I U.S. Department of Commerce Seattle, Washington Volume 97 Number 1 January 1999 The National Marine Fisheries Service (NMFS) does not approve, recommend, or endorse any proprietary product or proprietary material mentioned in this publication No reference shall be made to NMFS. or to this publication furnished by NMFS, in any advertising or sales promotion which would indicate or imply that NMFS approves, recommends, or endorses any proprietary product or proprietary material mentioned herein, or which has as its purpose an intent to cause directly or indirectly the adver- tised product to be used or purchased because of this NMFS publication. Fishery Bulletin JAN J 5 ms Contents Articles 1-8 Broadhurst, Matt K., Steven J. Kennelly, and Steve Eayrs Flow-related effects in prawn-trawl codends: potential for increasing the escape of unwanted fish through square-mesh panels 9-24 Brodziak, Jon, and Lisa Hendrickson An analysis of environmental effects on survey catches of squids Loligo pealei and ///ex illecebrosus in the northwest Atlantic 25-38 Chen, Yong, and Steve S. Montgomery Modeling the dynamics of eastern rock lobster, Jasus verreauxi, stock in New South Wales, Australia 39-52 Deree, Heather L. Age and growth, dietary habits, and parasitism of the fourbeard rockling, Enchelyopus cimbhus, from the Gulf of Maine 53-61 Heist, Edward J., and John R. Gold Genetic identification of sharks in the U.S. Atlantic large coastal shark fishery 62-70 Koeller, Peter Influence of temperature and effort on lobster catches at different temporal and spatial scales and the implications for stock assessments 71 -79 Lantry, Brian F., Donald J. Stewart, Peter S. Rand, and Edward L. Mills Evaluation of total-body electrical conductivity to estimate whole- body water content of yellow perch, Perca flavescens, and alewife, Alosa pseudoharengus 80-91 Limburg, Karin E., Michael L. Pace, and Kristin K. Arend Growth, mortality, and recruitment of larval Morone spp. in relation to food availability and temperature in the Hudson River 92-99 Lorenzo, Jose M., and Jose G. Pajuelo Biology of a deep benthopelagic fish, roudi escolar Promethichthys prometheus (Gempylidae), off the Canary Islands 100-109 Love, Milton S., and Korie Johnson Aspects of the life histories of grass rockfish, Sebastes rastreiiiger, and brown rockfish, 5. aurlculatus, from southern California Fishery Bulletin 97(1), 1999 110-117 McMillan, Charmion B. Three new species of hagfish (Myxinidae, Eptatretus) from the Galapagos Islands 118-131 Pollard, Morgan J., Michael J. Kingsford, and Stephen C. Battaglene Chemical marking of juvenile snapper, Pagrus auratus (Sparidae), by incorporation of strontium into dorsal spines 132-143 Polovina, Jeffrey J., Pierre Kleiber, and Donald R. Kobayashi Application of TOPEX-POSEIDON satellite altimetry to simulate transport dynamics of larvae of spiny lobster, Panulirus marginatus, in the Northw/estern Hawaiian Islands, 1993-1996 144-152 Poole, David, Geof H. Givens, and Adrian E. Raftery A proposed stock assessment method and its application to bowhead whales, Balaena mysticetus 153-169 Wintner, Sabine P., and Geremy Cliff Age and growth determination of the white shark, Carcharodon carcharlas, from the east coast of South Africa 170-184 Xiao, Yongshun, Lauren P. Brown, Terence I. Walker, and Andre E. Punt Estimation of instantaneous rates of tag shedding for school shark, Galeorhinus galeus, and gummy shark, Mustelus antarcticus, by conditional likelihood Notes 185-192 Kennelly, Steven J. Areas, depths, and times of high discard rates of scup, Stenotomus chrysops, during demersal fish trawling off the northeastern United States 193-199 Meyer, David L., Mark S. Fonseca, Patricia L. Murphey, Robert H. McMichael Jr., Michael M. Byerly, Michael W. LaCroix, Paula E. Whitfield, and Gordon W. Thayer Effects of live-bait shrimp trawling on seagrass beds and fish bycatch in Tampa Bay, Florida 200-211 Witzell, Wayne N. Distribution and relative abundance of sea turtles caught incidentally by the U.S. pelagic longline fleet in the western North Atlantic Ocean, 1992-1995 Abstract.— Two experiments were done in a flume tank to quantify the effects of codend mesh circumference and weight of catch on water flow. The mesh circumference of the posterior section of prawn-trawl codends (with and without bycatch reducing square- mesh panels) were assessed with three weights of catch (30 kg, 50 kg, and 70 kg). Compared with a codend with a cir- cumference of 100 meshes throughout its entire length, a codend with an an- terior section of 100 meshes and a pos- terior section of 200 meshes in circum- ference significantly increased the dis- placement of water forwards (up to 1120 mm from the end of the codend). This result varied with the weight of catch in the codend. The second experi- ment involved placing composite pan- els of square-shaped mesh (bycatch re- ducing devices) into the tops of the an- terior sections of two codends with the same configurations as those above (termed the "100 panel codend" and "200 panel codend," respectively). There was a displacement of water forwards immediately under the square-mesh panel in the 200 panel codend (by up to 2200 mm from the end of the codend). The results are discussed in terms of 1) the probable effects that codend mesh circumference and water dis- placement in codends have on fish be- havior, and 2) implications for the fu- ture development of bycatch reducing devices like square-mesh panels in prawn-trawls. Flow-related effects In prawn-trawl codends: potential for increasing the escape of unwanted fish through square-mesh panels Matt K. Broadhurst N.S.W, Fisheries Research Institute P.O. Box 21 Cronulla, New South Wales 2230, Australia Present address: Universidade Federal Rural de Pernambuco-UFRPE Departamento de Pesca, Laboratono de Oceanografia Pesqueira Av. Dom Manuel de Medeiros, s/n Dols Irmaos, Recife-PE, Brasil, CEP; 52 171-900 E-mail address: fhvhazin@truenet-Com.br Steven J. Kennelly N.S.W. Fisheries Research Institute PO. Box 21 Cronulla, New South Wales 2230, Australia Steve Eayrs Australian Maritime College PO. Box 21 Beaconsfield, Tasmania, 7270, Australia Manuscript accepted 30 April 1998. Fish. Bull. 97:1-8(1999). Oceanic prawn-trawling occurs from nine major ports in New South Wales (NSW) Australia and has a total value of approximately A$17 million per annum. The principal tar- get species is the eastern king prawn iPenaeus plebejus), although a signifi- cant proportion of the total value is derived from the sale of legally re- tained bycatch comprising individu- als of several species of fish, crusta- ceans, and cephalopods (see Kennelly, 1995). In addition to this landed catch, however, significant numbers of nontarget organisms of no commer- cial value are also captured and dis- carded, including juvenile fish of spe- cies which, when larger, are targeted in other commercial and recreational fisheries (Kennelly, 1995). Concerns over the incidental cap- ture and mortality of large numbers of juvenile fish have led to the de- velopment of various bycatch reduc- ing devices (BRDs), designed to minimize undesirable bycatch while maintaining catches of prawns and other commercially valuable indi- viduals (Broadhurst et al., 1996; Broadhurst and Kennelly, 1996; 1997). In particular, a new design comprising composite panels of square-shaped mesh (referred to as the composite square-mesh panel) was shown to increase catches of prawns (by up to 14% ) (Broadhurst and Kennelly, 1997) while signifi- cantly reducing up to 40% of total unwanted bycatch and up to 70% of the numbers of small individuals of commercially important species such as whiting iSillago spp.) (Broadhurst and Kennelly, 1996). The results from these experi- ments are attributed primarily to differences in the behavior of fish and prawns in the trawl. Fish are believed to be herded close together in the anterior section of the codend, upsetting the normal balance of the school and initiating an escape re- sponse towards the sides and top of the net and out through the open square-shaped meshes (Broadhurst and Kennelly, 1996). A contributing factor towards this escape is the dis- placement and change in direction of water flow due to the circumfer- Fishery Bulletin 97(1), 1999 ence of meshes in the posterior section of the codend. In an experiment to test this hypothesis, Broadhurst and Kennelly (1996) showed that a conventional codend made from an anterior section of 100 meshes circumference and a posterior section 200 meshes in circumference (a common commercial configuration) was less selective (i.e. retained more bycatch) than a conventional codend with a 100 mesh circumference throughout its entire length (see also Robertson and Stewart, 1988; Armstrong et al., 1990; Reeves et al., 1992; Galbraith et al., 1994). However, a compari- son of codends with the same netting configurations as above but containing composite square-mesh pan- els located in their anterior sections showed that the effects on selectivity due to increased circumference in the posterior section were negated by a signifi- cant increase in the escape of small fish (e.g. red spot whiting, Sillago flindersi) through the square meshes. These results led to the hypothesis that increased twine area and smaller mesh openings in codends with posterior sections of 200 meshes circumference increased the displacement of water forwards and out through the meshes in the anterior section (see also Watson, 1989). In turn, this water movement may have 1 ) physically directed small fish out through the strategically positioned composite square-mesh panel; 2) assisted them to maintain position in front of the catch in the codend, increasing their likelihood of ran- domly encountering square meshes; and (or) 3) stimu- lated their lateral line receptors and thus their overall escape response. The reaction of prawns to this stimuli was thought to be minimal, given their inability to sus- tain escape responses in trawls (see Lochhead, 1961; Newland and Chapman, 1989). Although the results from the paper discussed above led to several hypotheses about changes in water flow and fish behavior due to changes in codend geometry, we lacked the quantitative information on flow rates necessary to support or refute them. Such information is important for developing new designs of codends and understanding where to position square-mesh panels and other BRDs. Our goals in the present study were to quantify the effects on water flow at various positions in codends and un- der square-mesh panels in two flume tank experi- ments. We simulated commercial conditions in the flume tank by using different weights of catch in various codend designs. Materials and methods Two experiments were undertaken in May 1996 at the Australian Maritime College with the Faculty of Fisheries and Marine Environment's flume tank. This facility consists of a recirculating flow tank of fresh water, measuring 17.2 m long, 5 m wide, and 2.5 m deep and comprises three levels: 1) an upper level where nets, etc., are placed into the tank; 2) an observation level, with a continuous perspex view- ing-window; and 3) a water-return channel. The two lower levels feature a series of delivery bends and screens that maintain constant water velocity throughout the depth of the tank without any swirls or vortices. Several electric motors, hydraulic pumps, and impellor shafi-s provide water flow of up to 1.5 m/s. An electromagnetic current meter was attached to the base of a stainless steel stanchion (Fig. lA) and linked to a computer by means of a coaxial cable. The stanchion was attached to a movable carriage positioned on rails over the upper level of the flume tank. This assembly enabled the current meter to be repeatedly located at several predetermined positions within the tank. A full-scale Florida flyer prawn-trawl (material: 18 ply twine; mesh size: 40 mm) with a headline length of 5.4 m, was rigged to two fixed stanchions, located on the sides of the forward section of the flume tank. The trawl was rigged with a zipper (no. 10 ny- lon open-ended auto-lock plastic slides) to facilitate changing the codends. The codends used in the ex- periments were of normal commercial size and ma- terials, measuring 58 meshes long (2.3 m) and con- structed from 40-mm mesh netting and 60-ply UV- stabilized high-density polyethylene twine. These codends comprised two sections: the anterior section, which was 33 meshes long and attached to a zipper and the posterior section, which was 25 meshes long (for details see Fig. 1, B and C — see also Broadhurst and Kennelly 1996). Experiment 1 Two codend designs were compared. The codends (termed the 100 and 200 commercial codends) were made entirely of diamond-shaped meshes and com- prised anterior sections with a circumference of 100 meshes, attached to posterior sections with circum- ferences of 100 and 200 meshes, respectively (Fig. 1, B and C). Three incisions of the same size as the width of the current meter's stanchion (three meshes in length), were made in the tops of each codend at distances of 2200 mm, 1120 mm, and 560 mm for- ward ft^om the end of the codends to facilitate place- ment of the current meter inside the codends (Fig. IB). Experiment 2 The two codends compared in this experiment were similar to the 100 and 200 commercial codends de- Broadhurst et al.; Flow-related effects on prawn-trawl codends scribed above but included composite square-mesh panels made of 60-mm and 40-mm netting (3-ply and 48-ply UV- stabilized high-density poly- ethylene twine, respectively) cut on the bar and inserted into the tops of the anterior sections (termed the 100 panel and 200 panel codends) (for details see Fig. IC and Broad- hurst and Kennelly, 1996). Be- cause the current meter (and not its stanchion) was inserted only 5 cm into the tops of the codends (Fig. IC), it was not necessary to cut the meshes of these codends. Instead, four positions were labeled with a permanent marker at 2200 mm, 1720 mm, 1490 mm, and 1130 mm forward from the end of the codends (Fig. IC). Experimental procedure Thirty five rubber balloons were each filled with 2 liters of water, providing a total mass (in air) of 70 kg. These balloons were used to simulate masses of catch in the codends. In each experiment, the two codends were tested alter- nately. The particular codend to be examined was attached to the trawl and loaded ini- tially with 15 balloons (i.e. 30 kg). The hydraulic pumps in the flume tank were activated and adjusted to produce a flow of 1.2 m/s (the standard tow- ing speed during commercial operations). After a stabilizing period of 10 minutes, the stan- chion containing the current meter was alternately lowered into each of the predetermined positions in the codends (three in experiment 1 and four in experi- ment 2). After a further stabilizing period of one minute at each position, the current meter was switched on and left for a period of one minute, dur- ing which the flow of water immediately anterior to the current meter was recorded at one second inter- vals and the data were transmitted to the computer. 2,200 mm 2 kg Balloons 2,200 mm Figure 1 Diagrammatic representation of (A) the current meter and its stanchion, (B) 100 and 200 commercial codends and the three positions of water flow readings and (C) 100 and 200 panel codends and the four positions of water-flow readings. T = transver- sals; and N = normals. The mean flow rate from each minute of recording was calculated from these data and used in subse- quent analyses. After each reading, the current meter was moved to the next position and the procedure repeated. After six replicate readings were collected for each position, the flow of water in the flume tank was reduced to approx. 0.5 m/s and additional bal- loons were added to the trawl to simulate an increase Fishery Bulletin 97(1), 1999 in the volume of catch. The above procedure was then repeated to obtain data on the flow of water at each po- sition with 50 kg (25 bal- loons) and 70 kg (35 bal- loons) in each codend. Data analysis B>A>C>D I I 100 commercial codend H 200 commercial codend 2 . £ Water speed 11 . 1.0 09 S 08 07 06 30 Kg 50 Kg 70 Kg Position no. 1 Data collected in each ex- periment were examined by using Cochran's test for ho- mogeneity of variances and transformed if necessary. Data from experiment 1 were analyzed in a three- factor fully orthogonal, bal- anced analysis of variance. The factors were codends, positions, and weights. Data from experiment 2 were analyzed at each position in two-factor fully orthogonal, balanced analyses of vari- ance (Underwood, 1981). Significant differences de- tected in these analyses were investigated by Stu- dent-Newman-Keuls (SNK) multiple comparisons of means. Results Experiment 1 The analysis of variance showed that there were sig- nificant differences in flow rates between the type of codend, position of the current meter and weight in the codend, and significant interactions among codend-type, positions and weights (Table 1). SNK tests showed that mean water flow was greatest at position no. 2 (1.162 m/s) in the 100 commercial codend with a catch of 30 kg and lowest at position no. 3 (0.709 m/s) in the 200 commercial codend with 70 kg ( Fig. 2 ). SNK tests also showed that, compared with the 100 commercial codend, there was a signifi- cant reduction in mean water flow in the 200 com- mercial codend with 30 kg at position no. 2 (mean difference of 0.071 m/s) and across all three weights (30 kg, 50 kg, and 70 kg I at position no. 3 (mean dif- ferences of 0.203 m/s, 0.176 m/s and 0.184 m/s, re- spectively) (Fig. 2; Table 1). Although SNK tests did not detect differences in mean water flow between codends with 50 kg and 70 kg at position no. 2, nor across all weights at position no. 1, these combina- tions of weight and position showed trends similar In flume tank 1 30 Kg 50 Kg 70 Kg Position no. 2 30 Kg 50 Kg 70 Kg Position no. 3 Figure 2 Differences in mean flow rates ±SE between the 100 and 200 commercial codends tested in experiment 1 (< and > indicate direction of differences in SNK tests). Table 1 Summaries of F ratios from the analysis of variance to | determine effects on water flow due to different codends (i.e. 100 and 200 commercial codends), position of current | meter and weight in the codend. The square-root (sqrt(x)) transform was used to stabilize variances. In all tables | *P = <0.05;**P = <0.01. Treatment df Water How (m/s) Codends (C) 1 162.52" Position (P) 2 388.16** Weight (W) 2 62.42** C xP 2 35.01** C X W 2 8.04** p V W 4 12.39** C xPxW 4 0.81 Residual 90 to those described above — i.e. a mean reduction in water flow with the 200 commercial codend (Fig. 2). Experiment 2 There were significant differences in the mean wa- ter flow between the type of codend at position nos. 2, 3. and 4 and between weights at position nos. 2 and 3 (Table 2). There was a significant interaction between the type of codend and weight at position no. 1. SNK tests detected differences in the mean flow rates between the 100 and 200 panel codends at Broadhurst et al : Flow-related effects on prawn-trawl codends Summaries of F ratios (i.e. 100 panel vs. 200 Table 2 from analyses of variance to determine effects on water flow at different positions due to different codends panel) and weights. The transforms used to stabilize variances (if required) are also listed. Treatment df Position no. 1 Position no. 2 Position no. 3 sqrt(x) Position no. 4 Codend (C) Weight (W) C X W Residual 1 2 2 30 2.21 5.88* 2.29 3.95* 7.81** 0.87 5.16* 8.56** 0.74 7.92** 1.34 2.77 position nos. 1, 3, and 4 (Fig. 3). Compared with the 100 panel codend, there was significantly less water flow in the 200 panel codend with a catch of 70 kg at position no. 1 (0.115 m/s difference between means) and at position no. 3 with 50 kg (difference of 0.102 m/s) (Fig. 3, A and C; Table 2). Conversely, at posi- tion no. 4, SNK tests detected an increase in the mean water flow in the 200 panel codend with a catch of 30 kg, compared with the 100 panel codend (differ- ence of 0.067 m/s) (Fig. 3D). There was also a simi- lar, although not significant result for 50 kg at this position. There were no other significant differences, although, at position nos. 2 and 3, there were simi- lar trends across all weights (i.e. a reduction in wa- ter flow in the 200 panel codend) (Fig. 3, B and C). Discussion This study showed that the weight of catches and the configuration of the posterior section in codends can have significant effects on the displacement of water in the anterior section of codends. In inter- preting these results, it is important to note that under the simulated conditions in the present study, water was forced through a stationary trawl (at 1.2 m/s). Any localized displacements of water forwards in the codends examined in this study, therefore, can be expressed as a reduction in the flow entering the trawl and calculated by subtraction from 1.2 m/s. The flow of water at all positions in the four codends tested was less than 1.2 m/s, indicating that there was displacement of water anterior to the catch. However, the degree of this anterior water displace- ment varied significantly between the codends tested in each experiment (Figs. 2 and 3). For example, in experiment 1, the 200 commercial codend showed a significant increase (compared to the 100 commer- cial codend) in the displacement of water forwards at position no. 3 across all weights of catch (differ- ence in mean flow of up to 0.203 m/s [Fig. 2; Table 1]). These differences in flow may be attributed pri- marily to the distribution of the balloons used to simulate catch in the two codends and consequent changes in codend geometry. In the 200 commercial codend, the balloons were observed to spread out evenly in the posterior section, providing a greater surface area of catch incidental to the flow than in the 100 commercial codend. This effect, combined with the increase in the area of twine in the 200 com- mercial codend probably caused an increase in the displacement of water forwards in this codend. The above effects in the 200 commercial codend were also detected at position no. 2 with 30 kg of balloons and, although ANOVA failed to detect sig- nificant differences for the other weights at this po- sition, the trends in the results were similar (i.e. a reduction in flow in the 200 commercial codend com- pared with the 100 commercial codend [Fig. 2; Table 1] >. At position no. 1, for all weights, the force of the displaced water in front of the 200 commercial codend had dissipated to the extent where there were no sig- nificant differences between the two codends (Fig. 2; Table 1). It can be assumed, therefore, that the ma- jor influence of increased codend circumference on water displacement in the middle of the 200 commer- cial codend probably occurred up to some point between position nos. 2 and 3 (560 mm to 1120 mm from the end of the codend), in relation to the weight of catch. In experiment 2, the effects of increased codend circumference on water displacement under the square-mesh panel were detectable at a greater dis- tance from the end of the codend than those described above. Compared with the 100 panel codend, there were significant reductions in flow (corresponding to an increased displacement of water forwards) in the 200 panel codend at position nos. 1 and 3 (2200 mm and 1490 mm from end of the codend) with a weight of catch of 70 kg and 50 kg, respectively (mean dif- ferences in flow of 0.115 m/s and 0.102 m/s, respec- tively) (Fig. 3, A and C; Table 2). Although not sig- nificant, there were similar trends at position no. 2 for each weight and at position no. 3 for 30 kg and 70 kg (Fig. 3, B and C; Table 2). In contrast, there was a Fishery Bulletin 97(1), 1999 1.2 Position no. 1 09 08 12 B Position no. 2 •o c o B ^2-, C Position no. 3 significant increase in the flow of water at position no. 4 in the 200 panel codend with 30 kg and a similar result (though not statistically significant) for 50 kg (Fig. 3D; Table 2). This anomaly may be explained by the fact that the balloons in the 200 panel codend (like those in the 200 commercial codend) were orientated evenly across the surface area of the posterior section, increasing its diameter and decreasing the angle of incidence of its netting as it led into the ante- rior section of the codend at position no. 4. Although the current meter was located im- mediately under the compos- ite square-mesh panel, at this position it was effectively aligned slightly above the an- terior section of the codend and may have been influenced by the current outside the trawl, negating some of the flow-related effects of codend mesh circumference. With the exception of this latter result, the measured reductions in flow at most po- sitions in the 200 codends (cor- responding to increases in water displacement forwards) support the hypothesis that an increase in the circumference of meshes in the codend con- tributes towards the escape of small fish (between 5 and 20 cm) through the composite square-mesh panel (see Broad- hurst and Kennelly, 1996; 1997). The size of these fish suggests that they are using anaerobic muscle power to maintain position in the moving trawl (1.2 m/s) and are fatigued when they enter the codend (see Beamish, 1978; Wardle, 1989). A relatively small increase in the displacement of water forwards (e.g. 0.203 m/s at position no. 3 in the 200 commercial codend or 0.1 m/s at position no. 3 in the 200 panel codend I may be sufficient 1) to assist small fish to swim forwards and out through the square meshes in the panel and [or| 2) to enable them to reduce their tail-beat frequencies and main- Water speed in flume tank lllxili Water speed in flume tank llJill Water speed in flume tank LiJi 1.2 T^^ D^ositiormo. 4 Water speed in flume tank 1,1 - 08 iJilU 100 Panel 200 Panel 100 Panel 200 Panel 100 Panel 200 Panel 30 kg 50 kg 70 kg Figure 3 Differences in mean flow rates ±SE between the 100 and 200 panel codends tested in experiment 2 for each position of the current meter and for different weights {< and > indicate direction of differences in SNK tests of means). tain their position in the codend for a longer period, increasing their chances of random escape through the panel; and (or) 3) to stimulate their lateral line receptors and thus their overall escape. Without direct observations of fish swimming in the codend, it is difficult to determine their specific behavior during escape. Whatever their actual es- cape mechanism, however, the results obtained in this study provide important information for the sub- sequent design and location of BRDs like the com- Broadhurst et al : Flow-related effects on prawn-trawl codends Codend Composite square-mesh panel Semiporous panel of mesh Figure 4 Diagrammatic representation of proposed modification to codends ( taining the composite square-mesh panel. posite square-mesh panel. It is apparent that to maximize the effects of anteriorly displaced water in the posterior section of codends, the composite square-mesh panel should be located as close as pos- sible to the end of the codend, but sufficiently in front of the anticipated build-up of catch to prevent prawns from accumulating past the square meshes and escap- ing through them. A solution to this problem would be to increase the codend mesh circumference in the pos- terior panel (i.e. the 200 panel codend), causing the catch to spread laterally in the back of the codend, rather than to accumulate in front. Although such a modification would also increase surface area, displace- ment of water forwards, and probably enhance fish es- cape through the square-mesh panel, it would also re- duce mesh openings and the selectivity of the codend itself (Broadhurst and Kennelly, 1996). An alternative modification that may increase dis- placement of water forwards (other than increasing codend circumference and catch) is to move the com- posite square-mesh panel forward in the codend and create areas of "artificial catch" by using semiporous panels (e.g. those with fine mesh). These could be positioned on the bottom of the codend, behind the panel and at an angle to the direction of tow (e.g. see Fig. 4). Such modifications would produce similar flow-related effects as those observed in our study, i.e. they would displace water anteriorly, directing fish towards the panel. Future research into the re- finement of square-mesh panels and other BRDs in prawn-trawls that exploit the behavioral differences of fish in trawls, may benefit from these or similar modifications. Acknowledgments This work was funded by the Australian Fishing Industry Research and Develop- ment Council (grant no. 93/180). Thanks are extended to Bryan MacDonald, Garry Day, and Peter Cover for their assistance in conducting the experiments and to Takafumi Arimoto, John Watson, Jack Forrester, and Will Seidel for providing helpful discussions. Literature cited Armstrong, D. W., R. S. T. Ferro, D. N. MacLennan, and S. A. Reeves. 1990. Gear selectivity and the conservation of fish. J. Fish Biol. 37A:261-262. Beamish, F. W. H. 1978. Swimming capacity, /n W. S. Hoar and D. J. Randall (eds. ), Fish physiology, vol. VII: Locomotion, p. 101-187. Academic Press, New York, NY. Broadhurst, M. K. and S. J. Kennelly. 1996. Effects of the circumference of codends and a new design of square-mesh panel in reducing unwanted by-catch in the New South Wales oceanic prawn-trawl fishery, Australia. Fish. Res. 27:203-214. 1997. The composite square-mesh panel: a modification to codends for reducing unwanted bycatch and increasing catches of prawns throughout the New South Wales oce- anic prawn-trawl fishery. Fish. Bull. 95:653-664. Broadhurst, M. K., S. J. Kennelly, and G. O'Doherty. 1996. Effects of square-mesh panels in codends and of haulback delay on bycatch reduction in the oceanic prawn- trawl fishery of New South Wales, Australia. Fish. Bull. 94:412-422. Galbraith, R. D., R. J. Fryer, and S. M. S. Maitland. 1994. Demersal pair trawl cod-end selectivity models. Fish. Res. 20:13-27. Kennelly, S. J. 1995. The issue of bycatch in Australia's demersal trawl fisheries. Rev Fish Biol. Fish. 5:213-234. Lochhead, J. H. 1961. Locomotion, /n T H. Waterman (ed.). The physiol- ogy of the Crustacea, p. 313-356. Academic Press, New York, NY. Newland, P. L., and C. J. Chapman. 1989. The swimming and orientation behaviour of the Nor- way lobster, Nephrops norvegicus (L.), in relation to trawling. Fish. Res. 8:63-80. Reeves, S. A., D. W. Armstrong, R. J. Fryer, and K. A. Coull. 1992. The effects of mesh size, cod-end extension length and cod-end diameter on the selectivity of Scottish trawls and seines. ICES J. Mar Sci. 49:279-288. Robertson, J. H. B., and P. A. M. Stewart. 1988. A comparison of size selection of Haddock and Whit- Fishery Bulletin 97(1), 1999 ing by square and diamond mesh codends. J. Cons. Int. Explor. Mer 44:148-161. Underwood, A. J. 1981. Techniques of analysis of variance in experimental marine biology and ecology. Oceanogr. Mar. Biol. Ann. Rev. 19:513-605. Wardle, C. S. 1989. Understanding fish behaviour can lead to more se- lective fishing gears. In C. M. Campbell led.). Proceed- ings of the world symposium on fishing gear and fishing vessels, p. 12-18. Marine Institute, St Johns, Canada. Watson, J. W. 1989. Fish behaviour and trawl design: potential for selec- tive trawl development. In C. M. Campbell (ed.l. Proceed- ings of the world symposium on fishing gear and fishing vessels, p. 25-29. Marine Institute, St Johns, Canada. Abstract.— An analysis of environ- mental ei'fects on autumn survey catches of two commercially exploited squid species, Loligo pealei and Illcx illecchrosiis. was conducted. Research survey data collected during 1967-94 were used to determine the significance and relative importance of average depth of tow, time of day, bottom tem- perature, and surface temperature on bottom trawl catches of L. pealei. a ner- itic species, and /. illecebrosus. an oce- anic species. We examined habitat as- .■-(iciations of both species by using ran- domization methods and found that L. pealei was consistently associated with all of the environmental factors exam- ined. In comparison with L. pealei, catches of /. illecebrosus were much lower and associations with environ- mental factors were inconsistent. We also examined whether environmental conditions affected catches of juvenile and adult squid differentially. Depth had an important effect on the magni- tude of juvenile and adult L. pealei I catches, with the ratio of juvenile to ' adult catches decreasing with depth. Depth had a similar, but less pro- nounced, effect on /. illecebrosus catches. Time of day also affected L. pealei and /. illecebrosus catches. Catches of both species were lowest at night and diel effects were more pro- nounced for juveniles than for adults. Bottom and surface temperatures had a substantial effect on catches of juve- nile and adult L. pealei but had a vari- able influence on /. illecebrosus catches. The joint effects of depth stratification, time of day, and annual squid abun- dance on survey catches were also ana- lyzed to determine correction factors for diel differences in catchability of juve- nile and adult squid. Significant diel differences in catchability were de- tected for juvenile and adult L. pealei and for juvenile /. illecebrosus and diel correction factors were determined for survey catches of these size categories. In contrast, significant diel differences in catchability of adult /. illecebrosus were not detected. An analysis of environmental effects on survey catches of squids Loligo pealei and ///ex illecebrosus in the northwest Atlantic Jon Brodziak Northwest Fisheries Science Center National Marine Fisheries Service, NOAA 2030 S Marine Science Drive Newport, Oregon 97365-5296 E-mail address: |on.brodziak(a'noaa gov Lisa Hendrickson Northeast Fisheries Science Center National Marine Fisheries Service, NOAA 166 Water Street Woods Hole, Massachusetts 02543-1026 Manuscript accepted 19 March 1998. Fish. Bull, 97:9-24 (1999). An analysis of environmental ef- fects on survey catches of two com- mercially exploited squid species, Loligo pealei and Illex illecebrosus, was conducted for the continental shelf of the United States in the northwest Atlantic. Research sur- vey data, collected during autumn, were used to determine the signifi- cance and relative importance of average depth of tow, time of day, bottom temperature, and surface temperature on bottom trawl catches of L. pealei, a neritic species, and /. illecebrosus, an oceanic species. Both squids grow rapidly (Brodziak and Macy, 1996; Dawe and BeckM, appear to live less than one year (Hurley et al., 1985; Macy, 1995; Brodziak and Macy, 1996; Dawe and BeckM. and undertake seasonal migrations in response to fluctua- tions in food availability, water tem- perature, and spawning (Lange and Sissenwine, 1983; Rowell et al., 1985a; O'Dor and Coelho, 1993). The rapid growth and short lifespan of these sympatric species suggest that environmental conditions were probably important determinants of their distribution and abundance on the continental shelf. We investigated whether there was empirical evidence of habitat associations for both species by ap- plying the habitat association method of Perry and Smith (1994). This randomization method ac- counted for the stratified random sampling design of the Northeast Fisheries Science Center (NEFSC) survey and tested whether catches within a year were associated with depth, time of day, surface tempera- ture, and bottom temperature. Within-year associations that were consistent through time indicated which environmental conditions af- fected distribution and also pro- vided a qualitative indication of habitat preferences. We also examined whether envi- ronmental conditions affected catches of juvenile and adult squid differ- entially. Juveniles and adults of both species tend to differ in their food habits ( Vovk, 1972; Vinogradov and Noskov, 1979; Macy, 1982; Maurer and Bowman, 1985 ) and, as a result, may use different habitats for feeding. We compared the effects of average depth of tow, time of day. ' Dawe, E. G., and P. C. Beck. 1992. Popu- lation structure, growth, and sexual matu- ration of short-finned squid at Newfound- land, Canada, based on statolith analy- sis. ICES Council Meeting. Shellfish Committee/K, 3.3 p. 10 Fishery Bulletin 97(1), 1999 76 44 42 - 40 38 and surface and bottom tem- perature on survey catches to determine whether the habitat preferences of juveniles and adults overlap. Potential effects of different categories of depth, time of day, and surface tem- perature and bottom tempera- ture on mean catch per tow were tested for juveniles and adults of both species, and the potential impacts on the ratio of juvenile to adult catches were also examined. These analyses also suggested whether environ- mental effects were similar for these sympatric species, be- cause the degree of spatial over- lap between their geographic distributions is at a maximum during autumn. Last, we evaluated the rela- tive magnitude of diel effects on juvenile and adult squid catches in relation to survey design and fluctuations in annual squid abundance. Previous bottom trawl studies have shown that diel effects on catches of L. pealei and /. illecebrosus can be important (Roper and Young, 1975; Sissenwine and Bowman, 1978; Lange and Sissenwine, 1983; Arkhipkin and Fedulov, 1986; Shepherd and Forrester^) because greater catchability oc- curs during the day. As a result of diel vertical migrations, night catches of these two species can be biased low in relation to day catches. To account for diel effects on minimum swept-area estimates of L. pealei biom- ass and stock size, diel correction factors have been used to adjust nighttime bottom trawl catches to daytime equivalents (Lange and Sissenwine, 1980, 1983). However, these correction factors were devel- oped for total numbers of squid and were not size specific, even though Sissenwine and Bowman ( 1978) noted that a sixfold difference between the diel cor- rection factor for weight and numbers of L. pealei suggested differential vertical migration by size. In this study, we applied a general linear model to de- 74 72 70 68 66 36 Offshore depth strata surveys, 1963-94. Figure 1 for autumn Northeast Fisheries Science Center bottom trawl ^ Shepherd, G., and J. Forrester 1987. Diurnal variation in catchability during bottom trawl surveys off the Northeastern United States. ICES Council Meeting i987/B;44 ( mimeo), 15 p. termine size-specific diel correction factors for L. pealei and /. illecebrosus that accounted for poten- tial effects of survey design and fluctuations in an- nual abundance. When diel effects were significant, correction factors were determined to standardize nighttime catches to daytime units. Materials and methods Survey data Research survey data were analyzed from NEFSC autumn bottom trawl surveys conducted during 1967-94 between Cape Hatteras, North Carolina, and the Gulf of Maine. In general, autumn surveys were conducted from mid-September through mid- Brodziak and Hendnckson: Environmental effects on survey catcfies of Loltgo pealei and ///ex illecebrosus 11 October and along a similar cruise track each year. Standardized sui-vey gear, procedures, and the strati- fied random sampling design are described in Azarovitz ( 1981). Offshore survey strata are defined by four depth zones; ranging between 27 m and 366 m. Strata numbers 1-30, 33-40, and 61-76 were in- cluded in our analyses (Fig. 1). Sampling was con- ducted 24 hours a day. Autumn survey data were used because only in autumn are both squid species dis- tributed primarily within the survey sampling area. Survey data during 1963-66 could not be analyzed because catches of squid were not separated by spe- cies during these years. At each randomly selected survey station, sampling was performed with a no. 36 Yankee otter trawl rigged with roller gear. Only standard tows, consist- ing of 20-30 minutes duration at a vessel speed of 3.5 knots, were included in the analyses. After each tow, the total weight of L. pealei and /. illecebrosus was measured to the nearest 0.1 kg, each species was enumerated, and length-frequency data (mantle length [MLl measured to the nearest cm) were col- lected. The following hydrographic and navigational data were recorded for each station: depth at the start and end of every tow (m), time of tow (h:m, Eastern Standai-d Time ), bottom water temperature ( " C ), and surface water temperature (°C). Towing depth was computed as the average of depths recorded at the start and end of each tow. Surface water tempera- ture was included in this analysis because the effect of surface temperature on the vertical distribution and catchability of these squids was unknown but was anecdotally important and potentially different from the effect of bottom temperature during au- tumn. In general, bottom and surface water tempera- tures within the survey region vary during autumn owing to the transition from a stratified water col- umn, characteristic of summer, to a well-mixed con- dition typical of winter (Bowman, 1977). In particu- lar, the relationship between surface and bottom water temperatures varied on an annual basis dur- ing the 1967-94 autumn surveys. Surface and bot- tom temperatures were not significantly correlated in 8 out of 28 years (29'7c), and exhibited a positive correlation ( p =0.43 ) in 19 out of 28 years (68%) dur- ing 1967-94. When an environmental factor was not measured at a survey station, that station was excluded from the association test for that factor. Stations with missing environmental measurements occurred at random, with the exception of the Georges Bank por- tion of the autumn 1990 survey when no bottom tem- perature measurements were made. In total, 7177, 7179, 6105, and 6280 stations were available for the association tests of squid catch with depth, time of day, bottom temperature, and surface temperature, respectively. Within-year associations Perry and Smith ( 1994) developed a nonparametric test of association between an environmental factor and the quantity of catch during a stratified random survey. Their method uses the maximum absolute difference between the cumulative distribution func- tion (CDF) of an environmental factor and the catch- weighted CDF of that environmental factor as a test statistic in a randomization procedure to evaluate whether a significant association exists. In particular, the test algorithm is as follows. First, the empirical CDF if) of the environmental factor is computed as W, ni^=m.—^^^^nh (1) ^,h W, where t = the value of the environmental factor; /; = an index for the survey strata; i = an index for the tow in stratum /; ; the observed value of the environmen- tal factor from the /"' tow in stratum h ; ^ - the proportion of the sui'vey area in stra- tum li ; ri/^ - the number of tows in stratum h ; and KxJ- an indicator function with /(.v)=l, when X t. Second, the empirical cumulative distribution of catch as a function of the environmental factor (g) is computed as i:',"hys, (2) where and where y,^,= the catch from the i^^ tow in stra- tum h; y^^ = the mean catch in stratum /? ; and V . = the stratified mean catch. ^ St In Equation 2 above, the quotient y^^/y,, expresses the relative catch under environmental condition .r^^ in comparison with the stratified mean. The maxi- mum absolute value of the difference between f(t) and g(t) over all values of the environmental factor is the observed test statistic (T) where T = MAX y y ^/, I y,h - y^t lix,, (3) 12 Fishery Bulletin 97(1), 1999 To evaluate whether the test statistic is significant, the observed environmental measurements (.v^^ ) are randomly sampled with replacement and assigned to observed catches with probability W/^/rii^ under the hypothesis that the association between catch and the environmental factor is random. The value of the test statistic (T^l is then computed for this random assignment. The randomization procedure of assign- ing environmental measurements to catches and of computing the value of T^ is repeated a large num- ber of times to generate a distribution of test statis- tics for the null hypothesis of random association between catch and environmental factor. Last, the observed test statistic T is compared to the distribu- tion of test statistics T^ from the randomization pro- cedure to evaluate whether the null hypothesis of random association can be rejected. We applied this test to L. pealei and /. illecebrosiis catch, in numbers per tow, using four environmental factors: average depth of tow, time of tow, bottom water temperature, and surface water temperature. A total of 2000 randomizations were performed to provide an empirical distribution based on 2001 test statistics, including the original test statistic. Size-specific environmental effects Size-specific environmental effects on mean catches were evaluated separately for each species using tows that captured both juveniles and adults to ensure that comparisons were made under the same envi- ronmental conditions and that the ratio of juvenile to adult catch was well defined. Catch numbers of squid were separated into prerecruit (<8 cm ML for L. pealei and <10 cm ML for I. illecebrosus) and re- cruit (>8 cm ML for L. pealei and >10 cm ML for /. illecebrosus) size categories, where prerecruits and recruits roughly corresponded to juvenile and adult squid. In what follows, prerecruit and recruit cat- egories will sometimes be colloquially referred to as juveniles and adults, respectively. However, both L. pealei (Macy, 1980) and /. illecebrosus (Coelho and O'Dor, 1993) exhibit variability in sex-specific size at maturity, and for both species, an unmeasured fraction of smaller individuals within the recruit cat- egory were juveniles. The effect of depth on mean catch was evaluated first. Tow depth was categorized according to the depth zones used to define offshore strata of the NEFSC bottom trawl surveys: depth zone I (27-55 m), zone II (56-110 m), zone III (111- 185 m), and zone IV (186-366 m). Similarly time of tow was categorized into three time zones: zone I (night: 20:00-23:59 and 00:00 to 3:59), zone II (dawn and dusk: 4:00-7:59 and 16:00-19:59), and zone III (day: 8:00-15:59). Bottom and surface temperatures were grouped into three zones based on the 25th (P.,5) and 75th (P75) percentiles of the empirical tempera- ture distribution of each of these two variables. Tem- perature zone I was Pq to P.^j, zone II was P,,^ to P.^^, and zone III was P.^^ to Pj,,q. Mean catches between zones were compared after applying a logarithmic transformation to stabilize the variance of number per tow. Similarly, mean ratios of juvenile to adult catches were compared between zones after apply- ing a logarithmic transformation to the ratio. Mean logjg-transformed catch per tow values of juveniles and adults and their ratios were tested at the 5% level of significance by using the GT-2 test which is appropriate for unplanned comparisons with unequal sample sizes (Sokal and Rohlf, 1981). Only the ef- fects of individual environmental factors on juvenile and adult catches were evaluated in these univariate tests and the potentially important factors of interannual changes in abundance or of survey strati- fication were subsumed into random variation. Diel correction factors We analyzed the combined effects of survey design, time of day, and annual squid abundance to deter- mine correction factors for diel differences in catchability of juvenile and adult squid. As in the evaluation of size-specific environmental effects, catches (C) from tows that captured both juveniles and adults were logj^-transformed. A general linear model (Searle, 1987) was applied to estimate the ef- fects of survey stratum (Fig. 1), time of day category, and year on logj,, -transformed catches. Time of day was categorized into three time zones: zone I (night: 20:00-23:59 and 00:00-3:59), zone II (dawn and dusk: 4:00-7:59 and 16:00-19:59), and zone III (day: 8:00- 15:59) so that the time period effect (T) measured diel differences in mean catch. The year effect (Y) provided a measure of the effect of changes in rela- tive annual abundance on mean catch whereas the survey stratum effect (S) accounted for the effects of geographic location and depth. The general linear model (GLM) was log C,,^ = log C/fi + log T, + log Yj + log S^ + e,jk , (4 ) where C . - mean catch during the i"' time period (T, ) in the/'' year (7, ) within the k"' stratum (S^ ); Uj^ = mean catch in a standard reference cell (where standard time period=day. year=1994, and survey stratum=l ); and c , = an independent and identically distrib- uted normal random variable with zero mean and constant variance d^. Brodziak and Hendrlckson; Environmental effects on survey catches of Loligo pealei and ///ex illecebrosus 13 Parameters of the GLM model were estimated us- ing ANOVA for both prerecruit and recruit catches of each species. The significance of time, year, and stratum effects were evaluated by using type-Ill sums of squares that do not depend on the order in which effects are added to the model (Searle, 1987). Because daytime was the reference time period, the diel coefficient for the daytime had the value 1. Point estimates and confidence intervals of time zone KT, ) and II (T|j) coefficients in relation to the daytime co- efficient were computed whenever the diel effect was significant. Given estimates of Tj and Tjj, catches iy^/^ ) made during time zone I or II period can be corrected to daytime catch units by dividing them by the diel effects coefficient asy^/^ I Tj and y,,, / Tjj, respectively. Diel-standardized, stratified mean catches ( .y ,/■ ) can be expanded by the total number of sampling units within the survey region (N) to provide minimum swept-area estimates of juvenile or adult population size (Ny^^'') and variance (A^- Var[ v,,'- |) within the survey region using stratified random sampling es- timators (Cochran, 1977). Results Within-year associations The neritic squid species exhibited different degrees of within-year habitat associations than those exhib- ited by the oceanic species. Results of the random- ization tests of L. pealei catches with depth, time of day, bottom temperature, and surface temperature (Table 1) showed thatL. pealei catch was consistently associated with each of these factors. Associations were significant (P<0.05) in all years for depth, bot- tom temperature, and surface temperature. For time of day, associations were significant in all years ex- cept 1980 (96%). In contrast, results of the random- ization tests for/, illecebrosus (Table 2) indicated that catches of this species were inconsistently associated with only some of the factors. During the 28 years analyzed, a total of 15, 13, 7, and 12 associations were significant for depth, time of day, bottom tempera- ture, and surface temperature, respectively. All four factors were significantly associated with /. illecebrosus catch during 1983; otherwise, there was no apparent pattern of annual associations. Overall, L. pealei exhibited consistent within-year associa- tions, whereas /. illecebrosus exhibited variable within-year associations with the four environmen- tal factors. We compared the interquartile range of the catch- weighted CDF of each factor to the interquartile range of its unweighted CDF to see how associations Table 1 Results of univariate randomization test of association between catches of L. peak' and depth, time of day, bot- torn temperature , and surface temperature, during the 1 NEFSC autumn bottom trawl survey, 1967 -94. Table en- tries are the probabilities of random association' between L. pealei catches and the environmental factor. The sym- bol "*" denotes probability values of 0.05 > P> 0.01, and the symbol "**" denotes the probability values of 0.01 > P. Environmental factor Average Time Bottom Surface Year depth of day temperature temperature 1967 0.00** 0.00** 0.00** 0.00** 1968 0.00** 0.00** 0.00** 0.00** 1969 0.00** 0.00** 0.00** 0.00** 1970 0.00** 0.00** 0.00** 0.00** 1971 0.03* 0.00** 0.00** 0.00** 1972 0.00** 0.00** 0.00** 0.00** 197.3 0.04* 0.00** 0.00** 0.00** 1974 0.00** 0.00** 0.00** 0.00** 1975 0.00** 0.00** 0.00** 0.00** 1976 0.00** 0.00** 0.00** 0.00** 1977 0.00** 0.00** 0.00** 0.00** 1978 0.00** 0.00** 0.00** 0.00** 1979 0.00** 0.00** 0.00** 0.00** 1980 0.00** 0.07 0.00** 0.00** 1981 0.00** 0.00** 0.00** 0.00** 1982 0.01** 0.00** 0.03* 0.00** 1983 0.01** 0.00** 0.00** 0.00** 1984 0.00** 0.00** 0.00** 0.00** 1985 0.00** 0.00** 0.00** 0.00** 1986 0.01** 0.00** 0.00** 0.00** 1987 0.00** 0.00** 0.00** 0.00** 1988 0.00** 0.02* 0.00** 0.00** 1989 0.00** 0.00** 0.02* 0.00** 1990' 0.00** 0.00** 0.00** 0.00** 1991 0.00** 0.00** 0.00** 0.00** 1992 0.00** 0.00** 0.00** 0.00** 1993 0.02* 0.01** 0.00** 0.00** 1994 0.00** 0.00** 0.00** 0.00** ' No bot om temperature data were collected on Georges Bank daring 1990. varied across years. For brevity, the term "midrange" denotes the "average interquartile range" in what follows. For depth, this comparison of CDFs showed whether L. pealei or /. illecebrosus preferred shal- lower or deeper water in relation to observed depths. The midranges of depth for L. pealei and /. illecebrosus catches were 37-75 m and 79-149 m (Fig. 2A), respectively, whereas the midrange of all ob- served depths was 51-166 m. ForL. pealei, the aver- age of the median catch-weighted depth was 50 m, about 35 m shallower than the average observed depth and equal to the P2g of the observed depths. In contrast, the average of the median catch-weighted 14 Fishery Bulletin 97(1), 1999 depth for /. illecebrosus was 106 m, roughly 20 m deeper than the average observed depth. Overall, L. pealei was consistently associated with shallow depths (37-75 m), whereas /. illecebrosus was more common in deeper waters (79-149 m). The comparison of midranges for time of day indi- cated whether L. pealei or /. illecebrosus catches were more prevalent during night or day. We measured time in relation to a reference time of day (6:00 AM EST) because this roughly corresponds to first light, during autumn, when diel effects on the behavior of squid might be expected to change. This choice did not affect the results of the habitat association test; Table 2 Results of univariate randomization test of association | between catches of /. illecebrosus and depth , time of day, bottom tempera ture, surface temperature during the | NEFSC autumn bottom trawl survey, 1967- 94. Table en- tries are the probabilities of random associai ion' between /. illecet rosus catches and the environmental factor The symbol ' *" denotes probability values of 0.05 > P> 0.01, and the symbol "**" denotes the probability values of 0.0 1> P. Environmental factor Average Time Bottom Surface Year depth of day temperature emperature 1967 0.13 0.01** 0.01** 0.00** 1968 0.03* 0.24 0.05* 0.01** 1969 0.08 0.03* 0.26 0.33 1970 0.16 0.15 0.02* 0.02* 1971 0.01** 0.01** 0.00** 0.08 1972 0.24 0.30 0.12 0.03* 197.3 0.04* 0.13 0.71 0.12 1974 0.52 0.03* 0.03* 0.23 1975 0.34 0.00** 0.77 0.00** 1976 0.02* 0.20 0.54 0.30 1977 0.04* 0.44 0.26 0.01** 1978 0.12 0.00** 0.87 0.03* 1979 0.00** 0.15 0.07 0.07 1980 0.01** 0.01** 0.24 0.25 1981 0.61 0.01** 0.51 0.25 1982 0.03* 0.00** 0.09 0.18 1983 0.03* 0.00** 0.01** 0.00** 1984 0.03* 0.47 0.42 0.00** 198.5 0.00*' 0.00** 0.24 0.12 1986 0.03* 0.51 0.60 0.31 1987 0.39 0.34 0.53 0.04* 1988 0.98 0.12 0.06 0.34 1989 0.69 0.10 0.32 0.58 1990' 0.00** 0.00** 0.27 0.09 1991 0.04* 0.36 0.11 0.48 1992 0.13 0.01** 0.00** 0.00** 1993 0.33 0.24 0.53 0.03* 1994 0.04* 0.18 0.09 0.39 ' No bottom temperature data were collected on Georges Bank during 1990 however, median time of day was 18:00 EST, instead of 12:00 noon EST. The midranges for time of day for L. pealei and /. illecebrosus catches were 10:00-17:00 EST and 10:00-18:00 EST, respectively (Fig. 2B). In comparison, the midrange for time of day was 13:00- 1:00 EST. For both L. pealei and /. illecebrosus, the average median catch-weighted time of day was 14:00 EST, roughly 4 hours earlier than the average me- dian time of 18:00 EST, and roughly equal to Pgg of the overall time distribution. Overall, both L. pealei and /. illecebrosus exhibited diel catchability because squid catches were consistently greater during day than at night. Potential effects of bottom temperature were also examined to see whether L. pealei or I. illecebrosus preferred warmer or cooler bottom temperatures. The midranges of bottom temperature for L. pealei and /. illecebrosus catches were 11-15°C and 9-13°C, re- spectively, and the midrange for bottom temperature was 8-13°C (Fig. 2C). For L. pealei, the average of the median catch-weighted bottom temperature was 13°C, about 3°C warmer than the average of the over- all bottom temperature distribution. For I. illece- brosus, the average of the median catch-weighted bottom temperature was 11°C, which was nearly equal to the average of the observed bottom tempera- ture distribution Although L. pealei was associated with warmer bottom temperatures, it did not appear that /. illecebrosus was closely associated with bot- tom temperature. Effects of surface temperature were also examined to see whether L. pealei or /. illecebrosus preferred warmer or cooler surface temperatures. The mid- ranges of surface temperature for L. pealei and /. illecebrosus catches were 17-20°C and 13-20°C (Fig. 2D), respectively, and the midrange for all surface temperatures was 11-19°C. For L. pealei, the aver- age of the median catch-weighted surface tempera- ture was 18°C; about 4°C warmer than the average overall surface temperature distribution. Similarly, the median catch-weighted surface temperature of 16°C for /. illecebrosus was 2°C warmer than this average. Overall, L. pealei was generally associated with warmer surface temperatures, whereas /. illecebrosus associations were more variable, and this species was less frequently associated with warmer surface temperatures. Size-specific environmental effects Mean catches of L. pealei prerecruits and recruits varied across depth zones (Fig. 3A) where combined sample sizes for depth zones I, II, III, and IV were 1234, 1012, 270, and 99 tows, respectively. Mean catches of prerecruits peaked in zone I and declined Brodziak and Hendrickson: Environmental effects on survey catches of Loligo pealei and ///ex lllecebrosus 15 with depth, whereas mean catches of recruits peaked in zone III. The lowest mean catch of L. pealei prerecruits and recruits occurred in the deepest strata, zone IV. There was no signifi- cant difference between mean catches of prerecruits in depth zones II and III or recruits in zones I and II or zones II and III (Fig. 3Al. The mean ratio of prerecruit to recruit catches was highest in depth zone I (Fig. 3B) and ex- hibited a decHning trend with depth. However, the mean ratio was not significantly different in depth zones II, III, and IV. Overall, catches of L. pealei prerecruits decreased with depth, whereas recruit catches peaked at intermedi- ate depths (111-185 m). For /. illecebrosus, mean catches exhibited heterogeneity across depth zones (Fig, 3C), where sample sizes for depth zones I, II, III, and IV were 85, 283, 161, and 81 tows, respec- tively. Mean catches of recruits increased with depth and were highest in depth zone IV, whereas mean catches of prerecruits were rela- tively homogeneous across depths but were low- est in zone II. For prerecruits, mean catches were significantly different only between zones I and II (Fig. 3C). In contrast, mean catches of recruits were significantly different between all depth zones except zones III and IV and zones II and III. The mean ratio of prerecruit to re- cruit catches peaked in zone I, whereas the mean ratios in zones II, III, and IV were not significantly different (Fig. 3B). Overall, catches of/, illecebrosus prerecruits peaked in the shal- lowest depth zone (27-55 m) but were similar at greater depths. Similar to those of L. pealei. catches of/, illecebrosus recruits increased with depth as the ratio of prerecruit to recruit catch decreased with depth. As expected, mean catch of L. pealei prerecruits and recruits peaked during day and were low- est at night (Fig. 4A). Sample sizes for time zones I, II, and III were 518, 968, and 1 129 tows, respectively. For prerecruits, mean catch by time of day differed significantly for all three time zones (Fig. 4A), whereas mean catches of recruits were not significantly different between those for time zone II and zone III. The mean ratio of prerecruit to recruit catches peaked during day and was lowest at night (Fig. 4B), whereas the mean ratios for all time zones were significantly different (Fig. 4B). Overall, catches of both L. pealei prerecruits and recruits in- creased during day, although the difference be- tween day and night catches was more pro- nounced for prerecruits. 1967 1971 1975 1979 1983 1987 1991 1995 Midnight 6 PM- Noon 6AM 18 16 14 12 10 8 6 4 2 25 20 15 7. 10 1967 1971 1975 1979 1983 1987 1991 1995 1967 1971 1975 1979 1983 1987 1991 1995 Year 1967 1971 1975 1979 Year 1983 1987 1991 1995 Figure 2 Catch-weighted averages of (A) depth (m), (B) time of day (h), (C) bottom temperature (^Cl. and (D) surface temperature (°C) by year for L. pealei (solid circle) and/, illecebrosus (open square). The 25"', 50'^ and 75'^ percentiles (Pjj, P^o. and P,^ , thin solid lines) of the annual autumn distribution of each environmental factor are provided for comparison. 16 Fishen/ Bulletin 97(1), 1999 2.5 2,0 1.5 1.0 1,00 0.75 0.50 - f ° 25 - ra O — 0.00 ui o -' -0.25 -0.50 -0.75 -1.00 1.5 1.0 - 0.5 - Pre-recruits Recruits Jl <56m 56- 110m 111 -185m >185m B L. pealei I. illecebrosus I ¥"» 1 I T T <56m 56-110m 111-185m >185m Pre-recruits Recruits X JL mi X h U m_ <56m 56- 110m 111 - 185m Depth > 185 m Figure 3 (A) Mean catch of L. pealei. iBi mean ratio of prerecruit to recruit catches, and (C) mean catch of/, illecebrosus by depth for positive catches of L. pealei and /. illecebrosus prerecruits and recruits. For /. illecebrosus, the mean catch of prerecruits was highest during the day, whereas the mean catch of recruits peaked during dawn and dusk (Fig. 4C), where sample sizes for time zones I, II, and III were 86, 211, and 313 tows, respectively. Mean catches of prerecruits and recruits were lowest at night. For prerecruits, mean catches were not significantly different between time zones I and II (Fig. 4C). In contrast, mean catches of recruits were not significantly different between time zones II and III. Although the mean ratio of prerecruit to recruit catches peaked dur- ing the day ( Fig. 4B ), mean ratios were sig- nificantly different between time zones II and III. Similar to catches of L. pealei, catches of both size categories of /. illecebrosus were lowest at night. Mean catches of L. pealei prerecruits and recruits differed by bottom temperature zone (Fig. 5A) where temperature zones I, II, and III were 5.5 to 10.9°C (n=604), 10.9 to 16.1°C (n = l,340), and 16.1 to 28.0°C (n=272), respectively. Mean catch of prerecruits increased with bottom tem- perature and peaked in the warmest zone, whereas mean catch of recruits was great- est in zones II and III. In contrast, the mean catches of both size categories were lowest in temperature zone I. Mean catches of prerecruits were significantly different between all bottom temperature zones (Fig. 5A), whereas mean catches of recruits were not significantly different between zones II and III. The mean ratio of prerecruit to recruit catch peaked in zone III (Fig. 5B), and the mean ratios were not significantly different between zones I and II. Overall, catches of L. pealei prerecruits increased with increasing bot- tom temperature, whereas catches of re- cruits peaked at intermediate levels (11- 16^Cl. Catches of/, illecebrosus prerecruits and recruits also varied (Fig. SO with bottom temperature, where zones I, II, and III were 5.1 to 10.2°C (n = 122), 10.2 to 12.9°C (rt=242), and 12.9 to 25.5^C (n = 121), re- spectively. For /. illecebrosus prerecruits, mean catches were greatest in tempera- ture zones I and II, and the mean catch of recruits peaked in zone II. Mean catches of/, illecebrosus prerecruits and recruits were lowest in temperature zones I and III, Brodziak and Hendrickson: Environmental effects on survey catcties of Lollgo pealei and ///ex illecebrosus 17 respectively. For prerecruits, mean catches were significantly different only between temperature zones I and II (Fig. 5C), whereas mean catches of recruits were sig- nificantly different only between zones II and III. The mean ratio of prerecruit to recruit catches was greatest in tempera- ture zones II and III (Fig. 5D), however there was no significant difference be- tween the three zones. Overall, catches of /. illecebrosus prerecruits and recruits were relatively similar over the range of observed bottom temperatures (5-25°C). Mean catches of L. pealei prerecruits and recruits differed with surface tem- perature (Fig. 6A) where temperature zones I, II, and III were 7.1 to 14.8°C (n=615), 14.8 to 20.9°C (n=l,178), and 20.9 to 28.3°C (n=460), respectively Mean catches of prerecruits and recruits peaked in temperature zone II, whereas the low- est mean catch of prerecruits and recruits occurred in temperature zone I. Mean catches of both prerecruits and recruits were significantly different across all sur- face temperature zones (Fig. 6A). The mean ratio of prerecruit to recruit catches peaked in temperature zone II and the mean ratios were significantly different across all surface temperature zones (Fig. 6B). Overall, catches ofL. pealei prerecruits and recruits peaked at intermediate sur- face temperatures (15-21°C). Mean catches of/, illecebrosus prerecruits and recruits (Fig. 6C) also varied with sur- face temperature, where temperature zones I, II, and III were 7.8 to 14.4°C (/! = 123), 14.4 to 20.6°C («=242), and 20.6 to 26.5°C (n = 131), respectively. For /. illecebrosus prerecruits, mean catch peaked in temperature zone II but exhib- ited no trend across zones. In contrast, mean catch of recruits peaked in tempera- ture zone III and exhibited an increasing trend with surface temperature. There were no significant differences in mean catch of prerecruits, by surface tempera- ture zone (Fig. 6C), but mean catches of recruits were significantly different be- tween temperature zones I and III. The mean ratio of prerecruit to recruit catch peaked in temperature zone I (Fig. 6D) and declined with increasing surface tempera- ture. These mean ratios were significantly different between zones I and III. Over- 2.5 2.0 0.0 Pre-recruits Recruits DAWN or DUSK DAY NIGHT 1,5 ■^ 1.0 0.5 - 0.75 ■ 0.50 - B ^^ L, pealei 1 1 1. Illecebrosus x: o "to C) 0,25 - 0,00 - ■0,25 - -T 1 o _l 1 -0,50 - ^ "T -0,75 - y DAWN or DUSK DAY NIGHT ^^ Pre-recruits C 1 1 Recruits " j 1 1 L f 1 1 [ 1 1 1 1 DAWN or DUSK DAY Time of day NIGHT Figure 4 (Al Mean catch of L, pealei, (B) mean ratio of prerecruit to recruit catches, and (C) mean catch of/, illecebrosus by time of day for positive catches of L. pealei and /. illecebrosus prerecruits and recruits. 18 Fishery Bulletin 97(1), 1999 2.5 0.0 1.5 1.0 0.5 0.0 Pre-recruits Recruits 5.5- 10.9 10.9- 16.1 :10.9 10.9-16.1 >16.1 Pre-recruits Recruits T ji. 1 LiL JL, 5.1-10.2 10.2-12.9 12.9-25.5 Bottom temperature (C) Figure 5 (A) Mean catch of Z.. pealei. IB) mean ratio of L. pealei prerecruit to recruit catches, iC) mean catch of/, illecehrosus. and (D) mean ratio of /. illecebrosus prerecruit to recruit catches by bottom temperature for positive catches of prerecruits and recruits. all, catches of /. illecebrosus prerecruits were similar across the range of observed surface temperatures (8-26°C), whereas catches of recruits increased with surface temperature. Diel correction factors Diel effects from the GLM analyses were significant for L. pealei prerecruits and recruits and for /. illecebrosus prerecruits (Table 3). For L. pealei prerecruits, the time period effect was highly significant (F^.=280.79, P<0.001), and the year (F^,=6.05, P<0.001) and stratum effects (Fg=\&.23, P<0.001) were also significant. Estimates of the diel effects coefficients for L. pealei prerecruits were T[ = 0.0873 with 95% CI of [0.0713, 0.1068 ] and T„ = 0.4654 with 95%CI of [0.3958, 0.5472 ]. For com- parison, daytime catch rates of L. pealei prerecruits were roughly were about 11.5 times higher than time zone I values and 2.1 times higher than time zone II values. Similarly, for L. pealei recruits, the time period effect was highly significant (^5=100. 06, P<0.001), and the year (F,,=9.84, P<0.001 ) and stratum (^5^:12.45, P<0.001 ) effects were also significant. Esti- mates of the diel effects coefficients for L. pealei recruits were T, - 0.3420 with 95% CI of [0.2939, 0.3979 ] and Tj, = 0.8325 with 95%CI of [0.7372,0.9402 J.Time zone III catch rates of L. pealei recruits were roughly 2.9 times higher than time zone I values and about 1.2 times higher than zone II values. For/, illecebrosus prerecruits, the time period effect was significant (F^.=20.23, P<0.001 ). The stratum effect was also sig- nificant (Fg^l.98, P<0.001) but the year effect was not (^^=1.48, P=0.057). Diel ef- fects coefficients were estimated to be T, = 0.4251 with 95% CI of [0.3158, 0.5724 ] and T,i = 0.6281 with 95%CI of [0.5093, 0.7746 ]. By comparison, time zone III catch rates of /. illecebrosus prerecruits were about 2.4 times higher than zone I values and roughly 1.6 times higher than time zone II values. For /. illecebrosus re- cruits, the time period effect was not sig- nificant (Fg:^2.25, P=0.106), although the year (F^.=5.41, P<0.001) and stratum (P^.=3.05, P<0.001) effects were significant. For both species, diel effects were more pro- nounced for prerecruits than for recruits. Brodziak and Hendrickson: Environmental effects on survey catches of Loligo pealei and lllex lllecebrosus 19 o Discussion We found that L. pealei was consistently associated with all the environmental factors examined. The habitat associa- tions of L. pealei with depth, bottom tem- perature, and surface temperature indi- cated that these factors were important determinants of its autumn distribution. The consistent association of L. pealei with time of day appeared to be a conse- quence of the behavioral ecology of the species as it moves upward in the water column during the night to avoid preda- tion or to acquire prey. In comparison with L. pealei, autumn survey catches of /. illecebrosus were much lower and associations with envi- ronmental factors were inconsistent. For many of the years examined, /. illece- brosus catches were not associated with depth, temperature, or time of day. lllex illecebrosus feeds opportunistically in continen- tal shelf waters from Newfoundland to Cape Hatteras, during summer and autumn, prior to undertaking a lengthy offshore migration to spawning areas south of Cape Hatteras (O'Dor and Dawe, in press). Therefore, the lack of con- sistent habitat associations may be partly due to incomplete survey coverage of/, illecebrosus habitat during autumn and a lack of availabil- ity of this species to the bottom trawl survey gear. For example, the timing of the offshore and southward migration of/, illecebrosus may precede rather than follow the timing of the autumn survey in some years. In addition, few stations are sampled in the autumn habitat of this species, at the shoreward edge of the con- vergence zone, and these low sample sizes may make it difficult to detect habitat associations. Nevertheless, /. illecebrosus catches were sig- nificantly associated with depth, time of day, and surface temperature in approximately half of the years examined. Depth and surface tem- perature may be determinants of preferred habitat of /. Illecebrosus; however the impor- tance of these factors varied between years. The significant association of/, illecebrosus catches with daylight indicated that the species under- takes vertical migrations similar to those of L. pealei, but with less regularity as might be expected of an ommastrephid (Roper and Young, 1975). Depth had an important effect on the magnitude of juvenile and adult L. pealei catches. This associa- tion with depth corroborated previous studies. 0.50 c 10.2 10.2-12.9 >12.9 Bottom temperature (C) Figure 5 (continued) Table 3 Analysis of variance tables for general linear model results to estimate diel correction factors for L. pealei and /. illecebrosus prerecruits and recruits, where df is degrees of freedom, SS is type-III sums of squares, MS is mean square, F g is the F-statis- | tic, and P is the probabiUty value. Source of variation df SS MS Fs P L. pealei prerecruits Year 27 104.824 3.882 6.05 <0.001 Stratum .50 520.624 10.413 16.23 <0.001 Time of day 2 360.353 180.176 280.79 <0.001 L. pealei recruits Year 27 96.171 3.562 9.84 <0.001 Stratum 50 225.257 4.505 12.45 <0.001 Time of day 2 72.439 36.220 100.06 <0.001 /. illecebrosus prerecruits Year 27 9.761 0.362 1.48 0.057 Stratum 50 24.114 0.482 1.98 <0.001 Time of day 2 9.866 4.933 20.23 <0.001 /. illecebrosus recruits Year 27 41.351 1.532 5.41 <0.001 Stratum 50 43.274 0.865 3.05 <0.001 Time of day 2 1.275 0.637 2.25 0.106 Serchuk and Rathjen (1974) examined the distribu- tion and relative abundance of L. pealei and found that the highest catches were at depths less than 100 m during autumn. Vovk ( 1978) reported that the primary depth range of L. pealei was 50-100 m dur- ing September-November, and Lange and Sissen- 20 Fishery Bulletin 97(1), 1999 2.5 2.0 1.0 0.5 0.0 1.00 0.75 0,50 £" 0.25 o =i 0.00 d) o -' -025 •0.50 -0.75 -1.00 Pre-recruits Recruits 14.8-20.9 B < 14.8 14.8-20.9 >20.9 0.5 - 0.0 Pre-recruits Recruits 111 7.8-14 4 14.4-20.6 20.6-26.5 Surface temperature (C) Figure 6 (Al Mean catch of L. pealei. iB) mean ratio of Z,. pealei prerecruit to re- cruit catches, (C) mean catch of/, illecebrosus. and (D) mean ratio of/. illecebrosus prerecruit to recruit catches by .surface temperature for po.si- tive catches of prerecruits and recruits. wine (1983) reported relatively high catches at shallow (27-55 m) and also intermediate depths (111-185 m). In this study, however, depth affected catches of juvenile and adult L. pealei differently. Most shallow-water catches comprised juveniles, and this finding indicated that nearshore waters of the continental shelf constitute a preferred habitat of L. pealei juveniles during autumn. Catches of ju- veniles also decreased at greater depths. In contrast, catches of L. pealei adults peaked at depths near the edge of the continental shelf We detected a bathy- metric pattern of larger L. pealei with increasing depth. This pattern was also reported by Vovk (1978) on the basis of distant water fleet catches of L. pealei and is similar to the ontogenetic descent re- ported for some other loliginids (L. gahi, Hatfield et al., 1990; L. vulgaris reynaudii, Augustyn et al., 1992). In comparison with L. pealei, the effect of depth on /. illecebrosus was similar but less pronounced. This difference was probably due to the fact that /. illece- brosus utilize a wider range of depths than do L. pealei as they feed in shallow nearshore areas and undertake long dis- tance migrations in continental slope waters (O' Dor and Dawe, in press). None- theless, the empirical patterns of /. illecebrosus catches in relation to depth in our study area were consistent with those from different areas. In particular, Whitaker (1980) reported that /. illece- brosus catches were relatively low at depths less than 56 m and peaked at depths between 186 and 366 m in waters south of Cape Hatteras, whereas Grinkov and Rikhter'' reported that /. illecebrosus catches peaked at depths of 100-150 m along the edge of the continental shelf off Nova Scotia. In our study, much of the /. illecebrosus catch occurred at depths of roughly 80-150 m. As with L. pealei, we found that depth affected the catch of/. illecebrosus juveniles and adults differ- Grinkov, Y. A, and V. A. Rikhter. 1981. Some data on distribution of groundfish and short- finned squid along the oceanic slopes of the Scotian Shelf in spring. 1979. Northwest Atlan- tic Fisheries Organization (NAFO) SCR Doc. 81/ VI/63, sen no. N347, 13 p. I Brodziak and Hendrickson: Environmental effects on survey catches of Loligo pealei and lllex illecebmsus 21 ently. Catches of juveniles decreased at greater depths, and there was a bathy- metric pattern of larger /. illecebrosus with increasing depth. Although catches of adults increased with depth, in con- trast with L. pealei, catches of/, illece- brosus adults peaked at depths of 186- 366 m. This depth range corresponds to the convergence zone between continen- tal shelf and slope waters and intersects the shelf edge at roughly 150-200 m (Bowman, 1977). Commercial fishery catch data were consistent with this ob- servation, as Lange, Ingham, and Price* reported, noting that the highest catch rates recorded by domestic observers in the distant-water /. illecebrosus fishery were generally located within several miles of the shelf-slope front, at the shoreward edge of the convergence zone, between continental shelf and slope wa- ter. Overall, these observations suggested that the shelf-slope convergence zone is an impor- tant habitat for adult /. illecebrosus during autumn. Our analyses also demonstrated the importance of diel effects on L. pealei catches. This generally corroborated previous observations. Summers (1968) observed that L. pealei migrated vertically and that squid could be observed near the surface at night. Summers (1969) and Lange and Sissenwine (1983) observed that survey catches of L. pealei during the day were consistently higher than catches at night. Serchuk and Rathjen ( 1974) observed that 909^^ of L. pealei survey catches occurred during daylight, and Sissenwine and Bowman (1978) found that L. pealei catches were significantly higher during day. In our study, significant differences in diel effects were de- tected for both juveniles and adults. Catches of L. pealei juveniles and adults increased with increas- ing light availability although the diel effects were more pronounced for juveniles. In particular, time zone I and II catches of juveniles were 92% and 54% below mean catch during the day, whereas zone I and II catches of adults were 66% and 17% lower. The fact that the diel effect was more pronounced for ju- venile L. pealei may be related to differences in feed- ing behavior between juveniles and adults. L. pealei hatchlings must feed near the surface until their ten- tacles develop and they can capture larger prey -1.00 <14.4 14.4-20.6 >20.6 Surface temperature (C) Figure 6 (continued) ^ Lange. A.M. T.,M. C.Ingham, and C.A.Price. 1984. Distri- bution of maturing lllex illecebrosus relative to the shelf-slope water front of the northeastern United States. NAFO SCR Doc. 84/IX/109, Ser No. N906, 18 p. (Vecchione, 1981) whereas adults are primarily de- mersal and commonly rest on the bottom (Hanlon et al., 1983). Small juveniles feed primarily on crusta- ceans and gradually shift to a more diverse diet of crustaceans, fish, and squid as they grow and can capture larger and more energetically valuable prey (Vovk, 1972; Macy, 1982; Vovk, 1985; Anderson and GrLswold, 1988). As a result, juveniles need to un- dertake vertical migrations at night more frequently to capture a sufficient amount of suitable prey. The fact that /. illecebrosus catches were moder- ately associated with time of day was consistent with Sissenwine and Bowman (1978) and Shepherd and Forrester^ who found that time of day had an impor- tant effect on catches of/, illecebrosus. Diel vertical migrations may be related to /. illecebrosus feeding activity. Vinogradov and Noskov (1979) found that the feeding intensity of this species is greatest at night and lowest during the day. Similar to catches of L. pealei, catches of/, illecebrosus were higher during day, and diel effects were more important for juveniles than adults. In particular, catches of juve- niles at night and those at dawn and dusk were 37% and 57% below mean catch during the day. Guided by a review of the NEFSC domestic sea sampling program database, fishermen targeting/, illecebrosus take advantage of the behavior of this species by fish- ing only between dawn and dusk, when the squid are available to commercial bottom trawl gear. Schools of /. illecebrosus are not targeted at night because they are too dispersed near the surface of the water column. Instead schools of/, illecebrosus are targeted beginning at dawn, when they can be 22 Fishery Bulletin 97(1), 1999 consistently seen on sonar returning to the seabed.'^ In contrast, the ratios of juvenile to adult catches of /. illecebrosus were considerably lower than those for L. pealei. In part, this difference in ratios between species may reflect the importance of cannibalism for larger /. illecebrosus adults during autumn ( O'Dor and Dawe, in press). Although diel effects on I. illecebrosus juvenile catches were similar to L. pealei, they were less consistent. We found that bottom temperature had a signifi- cant effect on catches of L. pea/e; juveniles and adults. This generally corroborated prior studies. In particu- lar. Summers (1969) observed that large catches of L. pealei during winter were restricted to bottom tem- peratures of 8°C or higher. This lower temperature limitation was supported by Serchuk and Rathjen (1974) and Vovk( 1978) who reported that the major- ity of L. pealei catches during autumn occurred at bottom temperatures of roughly 9-14°C. Similarly, in our study, much of the L. pealei catch occurred at 11-15°C. However, catches ofL.pea/ez juveniles were highest when bottom temperatures exceeded 16°C, whereas catches of adults were highest when bot- tom temperatures were 11-16°C. Our results sug- gest that L. pealei juveniles generally prefer warmer bottom temperatures than do adults that appear to prefer intermediate bottom temperatures. This dif- ference in temperature preference might be expected if minor differences in temperature have a substan- tial impact upon growth rates of young squid (Forsythe, 1993). In contrast to bottom temperature, catches of L. peo/e; juveniles and adults had a similar pattern with respect to surface temperature. The highest catches of both L. pealei juveniles and adults occurred at tem- peratures of 15-2 1°C, whereas catches were lowest for temperatures below 15°C. Although L. pealei catches might be expected to increase with warmer water temperatures, adult catches peaked at inter- mediate bottom and surface temperatures and then declined at higher temperatures. This decline sug- gested that the higher temperature ranges observed in the survey were not optimal for L. pealei adults. Overall, the strong association of L. pealei with wa- ter temperature suggested that annual variation in patterns of ocean temperature affects the distribu- tion and influences growth and survival of this ner- itic species. In comparison to L. pealei, the less frequent asso- ciations of /. illecebrosus catches with bottom and surface temperatures suggested that temperature has a variable influence on the distribution of /. 5 Goodwin, G. 1997. Captain ofF/V Relentless and F/ V Per- sistence. Davisville, Rhode Island, 02882. Personal commun. illecebrosus from Cape Hatteras to the Gulf of Maine. Other studies generally supported the notion that /. illecebrosus are distributed over a broad range of tem- peratures. In particular, Whitaker (1980) reported that /. illecebrosus catches occurred over a wide range of bottom temperatures of 7-27°C, but that roughly 807( of the catch was taken in 8-10°C waters. Murawksi ( 1993) examined the mean latitudinal oc- currence of/, illecebrosus in relation to bottom and surface temperatures during autumn NEFSC bot- tom trawl surveys but found no statistically signifi- cant relationship. Rowell et al. ( 1985b) reported that /. illecebrosus appear to prefer bottom temperatures in excess of 6°C during summer on the Scotian Shelf, but that temperature did not appear to be a limiting factor In the present study, much of the/, illecebrosus catch occurred at bottom temperatures of 9-13°C and surface temperatures of 13-20°C. In comparison to L. pealei, I. illecebrosus appeared to prefer cooler bottom temperatures and surface temperatures. However, in contrast with L. pealei, bottom tempera- ture had a similar affect on catches of juvenile and adult /. illecebrosus. Surface temperature affected catches of juveniles and adults differently and, in particular, adult /. illecebrosus catches increased with surface temperature. Overall, /. illecebrosus catches occurred over a broader range of water temperatures in comparison with L. pealei, as might be expected of an oceanic species with a range that extends from temperate to boreal waters. Because the bottom trawl gear used on the NEFSC autumn survey only fishes 3.2 m above the seabed, diel differences in squid catches are expected when squid migrate vertically to acquire prey or to avoid predators. Significant differences were detected be- tween catches by time of day for both juvenile and adult L. pealei. As a consequence, it was inferred that diel correction factors were appropriate for survey catches of this neritic species. In contrast, diel ef- fects on survey catches of adult /. illecebrosus were not significant and diel correction factors were not developed for this oceanic species. The diel catcha- bility of L. pealei presents some challenges for the analysis and interpretation of bottom trawl survey data to estimate squid population totals. On one hand, diel correction factors developed in this study can be applied to time zone I and II catches to pro- vide a stratified mean catch per tow adjusted to day- light units. Resampling techniques, such as mirror- match bootstrapping, could be applied to estimate its variance (Smith, 1997) under appropriate distri- butional assumptions. In this case, all sur\'ey data could be used at the expense of additional computa- tional cost and potential bias induced by applying the diel correction factors to all survey strata. On Brodziak and Hendrickson: Environmental effects on survey catches of Loligo pealei and lllex illecebrosus 23 the other hand, squid population size and its vari- ance could be calculated on the basis of only day- light tows in the standard manner. In this case, roughly 1/3 of the available survey data would be used owing to the exclusion of time zone I and II tows. As a result, some survey strata would be under- sampled and precision would be lower. In general, the trade-off between bias due to diel correction and loss of precision due to use of only daylight tows warrants further study. Differences between L. pealei and /. illecebrosus catches by depth suggest that the stratification of the NEFSC bottom trawl survey, which was designed to sample groundfish populations, is appropriate for both species. Differences in L. pealei and /. illecebrosus catches by temperature imply that pre- ferred temperature ranges likely exist for both spe- cies within the survey region. However, temperature would not be a useful stratification variable for analy- sis of squid catches because temperature strata are dynamic and would fluctuate each year and because such a variable would lead to overstratification given the current depth and geographic stratification of the NEFSC survey. Regardless of how temperature af- fects survey catches of squids, the growth, recruit- ment, and abundance of L. pealei and /. illecebrosus can be expected to vary with ocean temperature re- gime (e.g. Dawe and Warren, 1993; Brodziak and Macy, 1996) because both are ecological opportun- ists with high, intrinsic population growth rates. Understanding how environmental factors, such as temperature, influence the productivity and distri- bution of squid stocks remains an important topic for fisheries research and management. Acknowledgments We thank the Captains, crew, and scientific staff of the RV Albatross TV and RV Delaware II for their dedicated efforts to survey fishery resources which provided the data for this study. We also thank F. Serchuk, L. Jacobson, and three anonymous reviewers for their helpful comments on the draft manuscript. Literature cited Anderson, J., and C. A. Griswold. 1988. The feeding ecology of juvenile long-finned squid. Loligo pealei, off the Northeast coast of the United States. U.S. Dep. Commer, NOAA, NMFS, MARMAP Contrib. FED/NEFSC 88-08, Narragansett, RI. 20 p. Arkhipkin, A. I., and P. P. Fedulov. 1986. Diel movements of juveniles lllex illecebrosus and other cephalopods in the shelf water-slope water frontal zone off the Scotian Shelf in spring. J. Northwest Atl. Fish. Sci. 7:15-24. Augustyn, C. J., M. R. Lipinski, and W. H. H. Sauer. 1992. Can the Loligo squid fishery be managed effectively? A synthesis of research on Loligo vulgaris reynaudii. In A. Payne, K. Brink, K. Mann, and R. Hilborn ( eds. ). Benguela trophic functioning, p. 903-918. S. Afr J. Mar Sci., Cape Town. Azarovitz, T. R. 1981. A brief historical review of the Woods Hole Labora- tory trawl survey time series. In W. G. Doubleday and D. Rivard (eds.). Bottom trawl surveys. Can. Spec. Sci. Pub. Fish. Aquat. Sci. 58:62-67. Bowman, M. J. 1977. Hydrographic properties. Marine Ecosystems Analy- sis [MESA] Program, MESA New York Bight Project. New York Sea Grant, Albany. NY, 78 p. Brodziak, J. K. T., and W. K. Macy III. 1996. Growth of long-finned squid, Loligo pealei. in the northwest Atlantic. Fish. Bull. 94:212-236. Cochran, W. G. 1977. Sampling techniques. 3'''' ed. John Wiley & Sons, New York, NY'. 428 p. Coelho, M. L., and R. K. O'Dor 1993. Maturation, spawning patterns, and mean size at maturity in the short-finned squid lllex illecebrosus. hi T. Okutani, R. K. O'Dor, and T. Kubodera (eds.). Recent advances in cephalopod fisheries biology, p. 81-91. Tokai L'niv. Press, Tokyo. Dawe, E. G., and W. G. Warren. 1993. Recruitment of short-finned squid in the Northwest Atlantic Ocean and some environmental relationships. J. Ceph. Biol. 2(21:1-21. Forsythe, J. W. 1993. A working hypothesis of how seasonal temperature change may impact the field growth of young cepha- lopods. In T. Okutani, R. K. O'Dor and T. Kubodera (eds. ), Recent advances in cephalopod fisheries biology, p. 133- 143. Tokai Univ. Press, Tokyo. Hanlon, R. T., R. F. Hixon, and W. H. Hulet. 1983. Survival, growth, and behavior of the loliginid squids, Loligo plei, Loligo pealei, and Lolliguncula brevis (Mol- lusca: Cephalopoda) in closed sea water systems. Biol. Bull. 165:637-685. Hatfield, E. M. C, P. G. Rodhouse, and J. Porebski. 1990. Demography and distribution of the Patagonian squid (Loligo gahi, d'Orbigny) during the austral winter. J. Cons. Cons. Int. Explor Mer 46:306-312. Hurley, G. V, P. Odense, R. K. O'Dor , and E. G. Dawe. 1985. Strontium labelling for verifying daily growth incre- ments in the statoliths of the short-finned squid, lllex illecebrosus. Can. J. Fish. Aquat. Sci. 42:380-383. Lange, A. M. T., and M. P. Sissenwine. 1980. Biological considerations relevant to the management of squid {Loligo pealei and lllex illecebrosus) of the north- west Atlantic. Mar Fish. Rev. 42 (7-8):23-38. 1983. Squid resources of the northwest Atlantic. In Ad- vances in assessment of world cephalopod resources, p. 21- 54. FAO Fish. Tech. Pap. 231, FAO, Rome. Macy, W. K., HI. 1980. The ecology of the common squid Loligo pealei (LeSueur), 1821. in Rhode Island waters. Ph.D. diss. Univ. Rhode Island, Narragansett, RI, 236 p. 1982. Feeding patterns of the long-finned squid, Loligo pealei, in New England waters. Biol. Bull. 162:28-38. 24 Fishery Bulletin 97(1), 1999 1995. Digital image processing to age long-finned squid using statoliths. In D. H. Secor, J. M. Dean, and S. E. Campana (eds.), Fish otolith research and application, p. 283-302. Univ. S. Carolina Press. Columbia. SC. Maurer, R. O., and R. E. Bowman. 1985. Food consumption of squids (///e.v lUecebrosus and Loligo pealei ) off the northeastern United States. NAFO Sci. Counc. Studies 9:117-124. Murawski, S. A. 1993. Climate change and marine fish distributions: fore- casting from historical analogy. Trans. Am. Fish. Soc. 122:647-658. O'Dor, R. K., and M. L. Coelho. 1993. Big squid, big currents, and big fisheries. In T. Okutani, R. K. O'Dor, and T. Kubodera (eds.). Recent ad- vances in cephalopod fisheries biology, p. 385-396. Tokai Univ. Press, Tokyo. O'Dor, R. K., and E. G. Dawe. In press. Illex illecebrosus. In R. K. O'Dor, P. G. Rodhouse, and E. G. Dawe (eds.). A review of the recruitment dynam- ics in the squid genus Illex. FAO Fish. Tech. Pap. Perry, R. I., and S. J. Smith. 1994. Identifying habitat associations of marine fishes us- ing survey data: and application to the northwest Atlantic. Can. J. Fish. Aquat. Sci. 51:589-602. Roper, C. F. E., and R. E. Young. 1975. Vertical distribution of pelagic cephalopods. Smithson. Contrib. Zool. 209, 51 p. Rowell, T. W., R. W. Trites, and E. G. Dawe. 1985a. Distribution of short-finned squid larvae and juve- niles in relation to the Gulf Stream frontal zone between Florida and Cape Hatteras. NAFO (Northwest Atlantic Fisheries Organization). Sci. Counc. Studies 9:77-92. Rowell, T. W., J. H. Young, J. C. Poulard, and J. P. Robin. 1985b. Changes in distribution and biological characteris- tics of Illex illecebrosus on the Scotian Shelf, 1980- 83. NAFO Sci. Counc. Studies 9:11-26. Searle, S. R. 1987. Linear models for unbalanced data. John Wiley & Sons, New York. NY, 536 p. Serchuk, F. M., and W. F. Rathjen. 1974. Aspects of the distribution and abundance of long- finned squid, Loligo pealei, between Cape Hatteras and Georges Bank. Mar. Fish. Rev 36(1):10-17. Sissenwine, M. P., and E. W. Bowman. 1978. An analysis of some factors affecting the catchability offish by bottom trawls. Int. Comm. Northwest Atl. Fish. Res. Bull. 13:81-87. Smith, S. J. 1997. Bootstrap confidence limits for groundfish trawl sur- vey estimates of mean abundance. Can. J. Fish. Aquat. Sci. 54:616-630. Sokal, R. R., and F. J. Rohlf. 1981. Biometry W.H. Freeman and Co., New York, NY. 859 p. Summers, W. C. 1968. The growth and size distribution of the current year c\ass Loligo peale,. Biol. Bull. 135(2):366-377. 1969. Winter population o{ Loligo pealei in the Mid-Atlan- tic Bight. Biol. Bull. 137( 1 ):202-216. Vecchione, M. 1981. Aspects of the early life history of Loligo pealei (Cephalopoda: Myopsida). .J. Shellfish Res. 1(2):171-180. Vinogradov, V. I., and A. S. Noskov. 1979. Feeding of short-finned squid, Illex illecebrosus, and long-finned squid, Loligo pealei, of Nova Scotia and New England, 1974-75. International Commission for the Northwest Atlantic Fisheries (ICNAF) Sel. Pap. 5:31-26. Vovk, A. N. 1972. Feeding habits of the North American squid, Loligo pealei Les. Trans. Atl. Sci. Res. Inst. Fish, and Oceanog. 42:141-151. (Canada Fish, and Mar. Sci. Transl. Ser 3304.] 1978. Peculiarities of the seasonal distribution of the North American squid Loligo pealei (Leseuer 1821 1. Malacol Rev. 11:130. 1985. Feeding spectrum of longfin squid iLoligo pealei) in the Northwest Atlantic and its position in the ecosystem. Northwest Atl. Fish. Org Sci. Counc. Studies 8:33-38. Whitaker, J. D. 1980. Squid catches resulting from trawl surveys off the southeastern United States. Mar Fish. Rev 42:39-43. r M ■1)1 19< abl ere 19i 18i Iwi ass pik 25 Abstract.— Rock lobster, Jasus ver- reauxi, have been fished off New South Wales, Australia, since the late nine- teenth century. Since 1994—95 (1 July 1994 to 30 June 1995) the fishery has been managed under an output-control scheme with an annual total allowable catch (TAC) of 106 metric tons (tl. Es- timates of catch and catch per unit of effort (CPUEi have been developed from data collected from the commer- cial fishery for the period 1903-1936 and the period from 1969-70 to 1993- 94. A production model was fitted to these data by using a robust observa- tion-error estimator that minimizes the median of squared differences between log-observed and predicted CPUEs. A bootstrap resampling procedure was incorporated into this robust estimation method to estimate stock parameters and their uncertainties. The virgin bio- mass of the rock lobster was 4084 t (its 5* and 95"^ percentiles being 2553 and 6400 t). The stock biomass decreased substantially until 1990-91. Since 1992-93, it has stabilized and has prob- ably increased owing to the large de- crease in the allowable catch after 1988-89. The stock biomass in year 1995-96 was likely to have been be- tween 15% and 30% of the virgin biom- ass (75% confidence interval). The im- plications from using different estima- tion methods on assessing this lobster stock are discussed. Modeling the dynamics of eastern tx>ck lobster, Jasus verreauxi, stock in New South Wales, Australia Yong Chen Steve S. Montgomery Fisheries Research Institute NSW Fisheries PO Box 21, Cronulia New South Wales 2230, Australia Present address (lor Y Chen): Fisheries and Marine Institute Memonal University of Newfoundland St. John's, Newfoundland, Canada A1C 5R3 E-mail address (for Y Chen) ychen a caribou ifmt nf ca Manuscript accepted 30 April 1998. Fish. Bull. 97:25-38(1999). The eastern rock lobster, Jasus ver- reauxi, reportedly is the largest spiny rock lobster in the world (Philips et al., 1980). It occurs in waters off the coast of New South Wales (NSW), Australia, around the coast of Tasmania, and as far west as South Australia (Montgomery, 1995). It is also found off New Zealand, predominantly around the North Island (Kensler, 1967a). Limited information on the life history of J. verreauxi in waters off NSW is available (McWilham and Philips, 1987; Montgomery, 1992, 1995; Montgomery and Kittaka, 1994; Montgomery^ ). Most informa- tion comes from studies of the spe- cies in waters off New Zealand (Kensler 1967a, 1967b, 1967c; Booth, 1984a, 1984b, 1986). How- ever, comparisons of mitochondrial DNA from juvenile rock lobsters from NSW and New Zealand waters have suggested that the populations are genetically distinct (Brasher et al., 1992). The distribution of rock lobsters across habitat is patchy. From the puerulus to early juvenile stages of their life cycle, lobsters are probably asocial and thought to occur princi- pally within the complex structure of forests of macroalgae or within beds of seagrass in waters from the intertidal zone to 30 m. During the older stages of the juvenile phase, the animal may begin to aggregate and migrate en masse to the habi- tat of adults. Adult rock lobsters live in aggregations from depths of around 10 m to those of the conti- nental slope (Montgomery, 1995; Montgomery^). From the older ju- venile stage onward, lobsters aggre- gate by day, and at night they roam alone. Information on the move- ments of tagged rock lobsters off New Zealand (Booth, 1984b) and spatial patterns in the length com- position of rock lobsters in waters off NSW (Montgomery^) suggest that for management purposes the entire NSW population of rock lob- sters should be considered as a unit stock. These studies indicate that older juveniles and adults move in an inshore-offshore direction and along the coast. The movement along the coast is thought to be as- sociated with breeding (Booth, 1984b). Rock lobsters have been fished off the east coast of Australia since the late nineteenth century. It is an important fishery in NSW, with an annual output of over 5 million US dollars. Since the 1994-95 fishing ' Montgomery, S. S. 1990. Preliminary study of the fishery for rock lobsters off the coast of New South Wales. Final Report, grant no. 86/64. Fisheries Research and Development Corporation, Canberra, Aus- tralia, 166 p. 26 Fishery Bulletin 97(1), 1999 year (i.e. from July 1 1994 to June 30 1995), this fish- ery has been managed under an output-control scheme with an annual total allowable catch (TAG) of 106 t. No attempt has been made previously to quantify the dynamics of this fishery. Abundance indices, catch per unit of effort (CPUE), have been developed by Montgomery ( 1995) from data collected in the commercial fishery for the period of 1903-36 and that fi-om 1969-70 (i.e. 1 July 1969 to 30 June 1970) to 1993-94. Because data are mainly col- lected and derived fi-om the NSW commercial fishery, large errors are likely to exist in the catch and CPUE data, and there is a concern that the quality of the data is perhaps not good for the purpose of modeling. Production models are fitted to catch and CPUE data by using an observation-error estimator that minimizes the sum of squared differences between log-observed and predicted CPUEs (Hilborn and Walters, 1992). This estimator assumes that there is only error in the observed abundance index and that there are no observation errors in catch or pro- cess errors in the dynamics of the stock biomass. Because the least-squares method is sensitive to the assumption on the error structure in the model (Rousseeuw and Leroy, 1987), the unrealistic error assumption associated with the observation-error estimator tends to result in large errors in estimated parameters when models are fitted to data (Schnute, 1989; Chen and Andrew, 1998). A more realistic error structure should include process error in the dynamics of the stock biomass and observation errors in both CPUE and catch. In our case, if the distribution of all error terms can be fully defined, we can apply the Kalman filter to gen- erate a likelihood function and then maximize this likelihood function to yield parameter estimates (Sullivan, 1992), or we can define an appropriate variance-covariance matrix based on the defined er- ror structure and then apply a generalized least- squares method to estimate parameters in the model (Paloheimo and Chen, 1996). However, the former approach is rather complicated because the dynamic model is nonlinear and there are two observation models (i.e. one for CPUE and the other for catch; Reed and Simons, 1996). The latter approach needs information on process and observation errors (Paloheimo and Chen, 1996). Such information is probably nonexistent in most fisheries. Moreover, the parametric assumption on error distribution (e.g. normal, log-normal, etc) may not be true. Because of all these difficulties with the error struc- ture for production models, it is desirable to have an estimation method that is robust with respect to as- sumptions concerning model error structure. Least median of squared errors (LMSE; Rousseeuw, 1984), which minimizes the median of squared differences between predicted and observed log CPUEs, is such an estimator (Chen and Andrew, 1998). A bootstrap procedure (Efron, 1979) was incorporated into the LMSE estimator to estimate the parameters and their uncertainties in this study. The probability of short-term overexploitation, defined as a fishing mortality rate higher than the selected biological reference points, was estimated for the next fishing season with different levels of TAC. Production models Production models are the simplest stock assessment models that are commonly used in fisheries (Hilborn and Walters, 1992). The input data for these models are the time series of catch and associated abundance index. Several variants of production models have been proposed (e.g. Pella and Tomlinson, 1969; Walters and Hilborn, 1976; Schnute, 1977, 1989; Punt, 1993). Without considering the structure of observation and process errors, the deterministic production model that is most commonly used can be written as B i^\ B.+g.-C, (1) where B c, s, = the stock biomass; - the catch; and = the growth of population in biomass, all in year i. The g^ is often referred to as surplus production and often described by the logistic or Schaefer function written as giB^) - rB^il - B/K), where r is a para- meter describing the intrinsic rate of population growth in biomass and AT is a parameter correspond- ing to the unfished equilibrium stock size (often re- ferred to as the carrying capacity or virgin biomass). The stock biomass in year ;' is often assumed to be directly related to a relative abundance index that can be observed in fisheries. This assumption can be written as where q - the catchability coefficient; and I^ = the abundance index in year i (Hilborn and Waters, 1992). Methods for the parameter estimation The use of an appropriate method to fit a production model to the observed data has been shown to be as Chen and Montgomery: Modeling the dynamics of Jasus verreauxi 27 important in terms of the reliability of estimated parameters as the specification of the algebraic form of the underlying population dynamic model (Punt, 1988, 1993; Polacheck et al., 1993). Several ap- proaches have been proposed to estimate parameters in production models when only indices of abundance and catch are available (Hilbom and Walters, 1992). The four most commonly used approaches are equilib- rium estimators (Gulland, 1961), effort-averaging es- timators (Fox, 1975), process-error estimators (Walters and Hilbom, 1976; Schnute, 1977), and observation- error estimators (Butterworth and Andrew, 1984; Ludwig and Walters, 1985). These approaches differ in how observation and process errors are introduced into the models that describe the dynamics of populations. Recently, it has been suggested in some studies that observation-error estimators tend to perform better than others in parameter estimation (Punt, 1988, 1993; Hilborn and Walters, 1992; Polacheck et al., 1993). These estimators are constructed by assum- ing that the population dynamic equations are de- terministic (thus there is no process error) and that all of the error occurs in the relationship between stock biomass and relative abundance index. This assumption can be written as log(/,) = \og{qB^ + e,. With the assumption that the e, are independent, normally distributed variates, the estimates of the model parameters (S^^^^^^^,, q, r, and K) are obtained by maximizing the appropriate likelihood function (Polacheck et al., 1993) or by minimizing the sum of squared e^ (Hilborn and Walters, 1992). The time series of stock biomass are estimated by projecting the biomass at the start of the catch series forward by using the historical annual catches. Because the estimation methods for observation-error estimators are least-squares types, they are sensitive to the as- sumption about the error structure of the model. Thus, parameter estimates tend to be unreliable if the specification of error structure (i.e. no process error, no error in observed catch, and log-normal er- ror in observed abundance index) is not correct. How- ever, in practice, it is almost impossible to know the true error structure. It is therefore desirable to use an estimation approach that is robust to the assump- tion about the model error structure for observation- error estimators. An observation-error estimator, which minimizes the median of squared differences between observed and predicted log CPUEs, has been found to be ro- bust with respect to incorrect specification of error structure (Chen and Andrew, 1998). This estimator can be written as Minimize median log(/,)-log(/,) / = l,...,iV It is an extension of the linear robust regression method used by Chen and Paloheimo (1994) and Chen et al.(1994) for nonlinear parameter estima- tion. It should be noted that the algorithm developed for the linear parameter estimation (Rousseeuw, 1984) can not be used for the above estimator. The simplex method of Nelder and Mead ( 1965 ) was used to conduct the nonlinear parameter estimation for the above estimator (Press et al., 1992; Chen and Andrew, 1998). Estimation of stock parameters for eastern rock lobster Catch and CPUE data were available for two time periods. The first period (hereafter referred to as period I) was from 1903 to 1936, and the second (pe- riod II) from 1969-70 to 1993-94. The fishery was confined to grounds close to shore in period I, whereas from the beginning of period II, the fishery expanded to the continental slope. Therefore, it is highly likely that large differences in catchability existed between these two periods. Both the size and structure of the rock lobster stock on the NSW coast may have changed greatly over the two periods (Montgomery, 1995), and it is likely that the growth rate of the NSW rock lobster stock differed between these two peri- ods of time. Parameters r and q were thus assumed to be different in these two time periods. Parameter K was assumed to be the same for these two time periods. This assumption was considered to be rea- sonable because the harvesting on the expanded fish- ing grounds at the beginning of period II was not from an unexploited portion of the stock. It is thought that eastern rock lobsters along the NSW coast dis- play a movement that is typical of several other spe- cies of rock lobster, moving between inshore and off- shore grounds and along the coast (see Herrnkind et al., 1994). Hence, lobsters on the grounds on the con- tinental slope likely had been exposed to fishing on more traditional shallower grounds at other times. The LMSE method was applied to fit the model to data observed in period I and estimate parameters ^1903' '"i' 9i' ^"^^ ^i' where subscript I refers to pe- riod I. Because the year 1903 was early in the devel- opment of the fishery, it is reasonable to assume that BjgQ3 is the same as K^ in parameter estimation (Hilborn and Walters, 1992). Thus, there are only three parameters to be estimated with data observed in the first time period. An algorithm that incorporates a bootstrap ap- proach into the LMSE method was developed to es- 28 Fishery Bulletin 97(1), 1999 timate the sampling distribution for model parameters. This approach is referred to as the bootstrapped LMSE method and its precedure fol- lows: 1 estimate the model parameters using the LMSE method; 2 calculate the predicted CPUEs using the LMSE-estimated pa- rameters; 3 calculate the residuals between observed and predicted log CPUEs; 4 randomly sample the residuals with replacement to add to the predicted logarithm CPUEs to yield pseudo observed CPUEs; 5 apply the LMSE algorithm to the pseudo observed CPUEs to esti- mate bootstrapped estimates; 6 repeat steps 4 to 5 100 times to simulate 100 sets of pseudo CPUE data and to estimate subsequently the corresponding 100 sets of boot- strapped parameters; and 7 calculate the median value and 90% confidence intervals for each parameter using the 100 boot- strapped estimates. 500 400 - 300 200 - 100 J^ 1900 1905 1910 1915 1920 1925 1930 1935 1940 (0 O 600 100 - -"S- Adjusted catch ~^ Reported catch 69-70 72-73 75-76 78-79 81-82 84-85 87-88 90-91 93-94 The landed lia, during .June 19701 Following Efron and Tibshirani ( 1985) and Manly ( 1991 ), the median values and 90% confidence intervals derived from the 100 sets of bootstrapped pa- rameters were used as the parameter estimates and their associated uncer- tainties. The 100 bootstrap runs were considered sufficient for the LMSE method in this study because a preliminary analysis indicated that differ- ences in the distributions of estimated parameters de- rived from 100 and 2000 bootstrap runs were small. Since K^^ was assumed to be the same as /C,, there were three parameters, Tj,, (Jjj, and Sgg y^, to be esti- mated in modelling CPUE and catch data observed in period 11. Although CPUE for period 11 was de- rived from accurate records of catch and effort asso- ciated with part of the commercial fishery, under- reporting of total catch occurred to a significant ex- tent during period II (Montgomery and Chen, 1996). The extent of under-reporting in different fishing seasons was estimated on the basis of a survey of fishermen (Table 1; Montgomery'; Montgomery, unpubl. data). Thus, catch in fishing year J was ad- justed from the reported catch (Fig. 1) as Fishing year Figure 1 catch of eastern rock lobsters in New South Wales (NSW), Austra- the period of 1903-36, and from 1969-70 (i.e. 1 July 1969 to 30 to 1993-94. {reported catch) j ( Adjusted catch) . = 1 - (adjustment coefficient } J where adjustment coefficient = the proportion of der-reporting estimated. un- The seven-step procedure described above was modified to include the uncertainties in K^ in esti- mating the parameters and associated uncertainties in period II. The modification was accomplished by using values of K^ randomly sampled from the bootstrapped samples generated in the bootstrapped LMSE analysis for period I. Such a modification takes into consideration the variation in K^ (thus K^^) when estimating q,,, r„, and Bjgyg ^^and their variabilities. Chen and Montgomery: Modeling the dynamics of Jasus veireauxi 29 Table 1 Coefficients used to adjust reported catch data in the pe- riod of 1969-70 to 1993-94. Year Adjustment coefficient 1969-70 to 1979-80 1980-81 to 1989-90 1990-91 to 1991-92 1992-93 to 1993-94 0.5 0.7 0.45 0.15 Evaluation of probability of overexploitation for next fishing season The probability of short-term overexploitation (i.e. the probability of the fishing mortality rate being higher than defined biological reference points) was defined with respect to different levels of catch for the 1996-97 fishing season. The two biological refer- ence points used in our study were /], j and Ay<,y- '^^^ /Ji J is equivalent to the more commonly used F,, j (Hilborn and Walters, 1992) and is defined by the equation dC{E) dE = 0.1 dC(E] E=t„,.q dE E=0 where C(E) - the equilibrium yield corresponding to efforts (Punt, 1993). From the above equation and Equation 1, /"o j can be calculated as/"„ , = 0.45/-. The TAG in year,/ based on the/j^i , can be calculated as TAC^ ^(j) - 0A5rB where B is the estimate of stock biomass in yearj. The rate of fishing mortality 'fv/sy' producing maximum sus- tainable yield (MSy) can be calculated as/'^^^^.y. = 0.5/-. The TAC in year j, based on /"^^^y, can be calculated as TACi^fgy ij) = 0.5rB^. It should be noted that the TAC^jgY 0' calculated above changes with the cur- rent stock biomass and is thus dynamic over time. It differs from the commonly used equilibrium maxi- mum sustainable yield (EMSY) calculated as rK/4 (Hilborn and Walters, 1992). The use of TAC MSY C7» is more robust with respect to interannual variabil- ity in the biomass of the stock than is a TAC based on the more commonly used EMSY. Results ,, (7,, and K^ estimated with the LMSE method were 0.203, 1.76 x lO*^ ' Parameters '"i , , (per vessel, and 3208 1, respectively This results in an EMSY of 163 t. The predicted CPUEs tended to follow the CPUEs ob- Table 2 Summary of the estimates of parameters with the bootstrapped LMSE method from 100 runs of bootstrap simulation for CPUE and catch data observed during 1903 to 1936. Parameter Statistic '■l (?[ X 10-6 (per vessel) K, a' Median Mean CV 5th% 95th% 0.155 0.149 48.6'* 0.045 0.292 1.20 1.26 28.0% 0.79 1.94 4084 4269 26.0'7<- 2554 6400 0.428 0.497 53.1% 0.187 1.058 ' For each bootstrap run. a is calculated as n ,7 = 1^ CPUBj-CPUEf where CPUE and CPUE are observed and predicted catch per unit of effort, respectively, and n is the number of years. served in the majority of years in period I (Fig. 2). However, CPUEs observed in the years 1904 and 1917-24 differed considerably from the LMSE-pre- dicted values, indicating that they virtually were ig- nored in the parameter estimation. The estimated stock size in year 1936 was 37. 6^^ of the virgin biom- ass. The median value for K^ from the 100 boot- strapped LMSE estimates was 4084 1 (Table 2), about 27*7^ higher than the LMSE-estimated K^. The coef ficient of variation (CV) for the LMSE-estimated K^ was only 269f , indicating that the uncertainty asso- ciated with the LMSE-estimated K^ was small. The median values of bootstrapped r and q were 0.155 and 1.2 x lO"*^ (per vessel) (Table 2), about 24% and 32% lower than the LMSE-estimated r and g, respec- tively. The higher CV for /-, compared with the CVs for q and K^, indicates that the estimate of r is more uncertain than the estimates of g and K^. The distri- butions of all three parameters tended to be posi- tively skewed (Fig. 3). The estimated EMSY ranged from 50 t to 200 t with the median value of 151 t (Fig. 3). By assuming K^^ to be the same as the median value of if J estimated in the bootstrapped LMSE estimation, we estimated parameters rjj, Qjj, and 5i9g9_7y The LMSE-estimated Sjggg_.yg was close to the value of Xj (4,365 t). The r„ was 0.218, about 7% higher than the LMSE-estimated r,. The LMSE-estimated gjj was 0. 15 X 10"*' ( per trap-month ). The model was fitted by means of the LMSE method by ignoring the data observed in the two fishing seasons 1971-72 and 1974-75 because 30 Fishery Bulletin 97(1), 1999 "5) UJ O 0 — 1900 ~B~ Observed cpue LMSE 1905 1910 1915 1920 Year 1925 1930 1935 1940 Figure 2 Catch per unit of effort observed and predicted by the LMSE method for the NSW eastern rock lobster fishery in the period of 1903 to 1936. CPUEs in these two fishing seasons increased greatly and abruptly from the previous years, followed by an equally abrupt decrease (Fig. 4). Distributions for parameters of the model estimated with the bootstrapped LMSE method for pe- riod II are presented in Figure 5. The median value of the stock bio- mass in year 1969-70 was 3808 t, 13% lower than the corresponding LMSE estimate. The estimated median value of Tj, was 0.172 (Table 3), 227f lower than the LMSE-estimated Tjj. The median value of (7jj was only 10% of that estimated for Qj. This resulted from different units of fishing ef- forts used in calculating the abun- dance index (CPUE) in the two O.IO o.os o.oa 1 0.04 002 0.00 h II 1 .ll J 2.000 2.800 3 400 4,000 4.600 5,200 6,600 6 400 7, COO 7,600 K(t) m o o.ie 0.12 0.08 III 1 0.04 .1 nil IIIlIi^ 0 00 0 06 0.12 0 18 0 24 0.30 0 36 60 80 110 140 170 200 q X 10°(1 /vessel) Maximum sustainable yield (t) Figure 3 The distributions of the parameters estimated with the bootstrapped least median of squared errors (LMSE) for the period of 1903 to 1936. Chen and Montgomery: Modeling the dynamics of Jasus verreauxi 31 1.2 time periods (per vessel versus per trap-month). The CVs and 90% confidence intervals for the parameters were small (Table 3), indicating that esti- mates of parameters for period II had small uncertainties. The estimated EMSY ranged from 80 t to over 200 t with a median value of 120 t (Fig. 5). The median and 5'*^ and 95* percentiles of the biomass of the stock for each year in pe- riod II are plotted in Figure 6. The calculation of total catches for 1994-95 and 1995-96 had not been completed when this study was conducted. How- ever, the TAG of 106 t was likely to have been fully real- ized in each of these years. Thus, the biomasses of the stock for 1995-96 and 1996- 97 were projected based upon the assumption that catches in 1994-95 and 1995- 96 were 106 t. The 90% confidence intervals tended to increase from 1969-70 to 1994-95 (Fig. 6), indi- cating that estimates of the stock biomass in recent years were less precise than those in earlier years. The median value of the biomass of stock in the 1995- 96 fishing year was 1420 t, and the 5'^ and 95"^ per- Observed cpus LMSE \ i i i h 0.0 1389-70 72-73 7S-7S i I i I I i i I I I I I I I I I 78-79 81-82 84-85 87-88 90-91 93-94 Fishing year Figure 4 Catch per unit of effort observed and predicted with the LMSE method for the NSW eastern rock lobster fishery during the period of 1969-70 (i.e. 1 July 1969 to 30 June 1970) to 1993-94 (i.e. 1 July 1993 to 30 June 1994). Table 3 Summary of the estimates of parameters using the LMSE method from 100 runs of bootstrap simulation for CPUE and catch data observed from 1969-70 to 1993-94. Parameter Statistic '•ll <7„xlO-* (trap X month) ' "1969-70 (ton) a' Median 0.172 0.15 3808 0.237 Mean 0.177 0.15 3764 0.241 CV 19.4% 16.2% 11.5% 360% 5"'% 0.132 0.11 3099 0.059 gs"-* 0.223 0.18 4418 1.098 ' For each bootstrap -un a is calculated as y, CPUE J - CPUE J )^ ,T-1 j=i where CPUE and CPUE are observed and predicted catch per unit of effort, respectively, and n is the number of years. centiles were 710 and 2719 t (Fig. 6). Median values for the biomass of the stock fell until 1990-91 and then started to increase from 1992-93. The plot of the distribution of the ratio between biomass in year 1995-96 against the biomass of the virgin stock in- dicated that there was more than 75% chance that the biomass ranged between 15% and 30% of the vir- gin biomass (Fig. 7). The distribution of the biomass of the stock in 1996-97 is summarized in Figure 8. There was more than a 70% chance that the biomass of the stock ranged between 1000 to 1750 t. The probabilities of short-term overharvest (i.e. exceeding the selected biological reference points) were calculated for dif- ferent levels of catch in 1996-97 on the basis of the distribution estimated for the biomass of the stock in 1996-97 (Fig. 9). For example, the TAG of 100 t would have a 30% chance of exceeding the reference point f^^Y' ^^'^ ^ ^^0% chance of exceeding /q ^ (Fig. 9). Discussion The model used in this study is one of the simplest models commonly used in fish stock assessment (Hilborn and Walters, 1992). An essential assump- tion of this model, as with similar models, is that the relationship between GPUE and the biomass of the stock remains constant over time. Because of the use of the robust estimator, this assumption becomes the following: the proportional relationship between the 32 Fishery Bulletin 97(1), 1999 (0 XI o 0.1 0.12 014 0.16 0.18 02 0.22 "0.24 0.06 008 0.1 012 0.14 0.16 0 18 0.2 0 22 0.24 0.26 q X 10^[1/trapx month)] 2.000 2,600 3.000 3,500 4.000 4 500 5.000 Stock biomass in year 1969-70 (t) Maximum sustainable yield (t) Figure 5 The distributions of the parameters estimated with the bootstrapped least median of squared errors (LMSE) for the period of 1969-70 to 1993-94. stock biomass and CPUE is the same for the major- ity of years. Compared with traditional least-squares estima- tion methods, the LMSE method tends to fit the model to the majority of the data, and its estimates of parameters are not affected by atypical data ob- served in a few years (Chen et al., 1994). Thus, if atypical observations arise, parameter estimation with the LMSE will not be affected greatly. Because the LMSE is not sensitive to atypical data, these data tend to be far from the LMSE-estimated CPUEs (Figs. 2 and 3) and thus are readily detected. How- ever, it is important to determine why an observed CPUE is far from the predicted CPUE. Such a prac- tice requires extensive background information on the fishery. Apart from exceptionally large observa- tion errors, atypical observations may arise from unusual environmental conditions and substantial changes in fishing methods and locations. The na- ture of atypical observations resulting from unusual environmental conditions differs from those result- ing from unusual observation errors. These atypical data should be considered separately from those ob- served under normal conditions in the stock assess- ment process because such unusual conditions last only for a short period of time in the development of the fishery (Chen et al., 1994). For example, the ob- served CPUE in 1971-72 was much higher than the LMSE-estimated CPUE (Fig. 4). This might result from the fishery being expanded in that year to in- clude aggregations of lobsters on the previously unexploited slopes of the continental shelf. This would result in a high q for that year. Clearly, such a high level of CPUE cannot be sustained for long and should not be interpreted as an indicator of large stock biomass for that year CPUEs observed from years 1917 to 1924 were much lower than the LMSE predicted ( Fig. 2 ). This finding may be related to large observation errors in CPUEs. For this period, the number of fishing boats (other than trawlers or Dan- Chen and Montgomery: Modeling the dynamics of Jasus verreauxi 33 ish seiners) that operated from lobster producing ports and in oceanic waters was used as an index of fishing effort for the lobster fishery. This was done because it was impossible to distinguish between vessels that were used for fishing lob- sters and vessels that were used for fishing other species in the same area (Montgomery, 1995). During this period, an- nual landings of lobsters were probably under-reported be- cause only lobsters sold at the main market centers were re- corded. Such an over-estima- tion of efforts and under-esti- mation of catches would lead to under-estimation of CPUE for this period of time. It should be realized, however, that the above reasons for the large deviation of observed data from model estimates re- main a working hypothesis. Outliers can be identified in linear regression analyses by using criteria developed by Rousseeuw and Leroy (1987). However, the criteria devel- oped through extensive simu- lations based on a linear re- gression model cannot be used in nonlinear regression analy- sis. An extensive simulation study is needed to develop suit- able criteria to identify outli- ers in nonlinear regression analyses. Before this can be done, the LMSE-based re- weighted least squares, which has been shown to behave bet- ter than the least median of squares in linear regression analyses (Chen et al., 1994; Chen and Paloheimo, 1995), cannot be used in nonlinear regression analyses. The results presented in our paper were based on catch data adjusted by using one set of adjustment coefficients. To test the sensitivity of the results to the adjustment coefficients, we also used two other sets of adjustment coefficients in the analysis (Table 4). These two sets of coefficients have more extreme values than the one presented in Table 1. The detailed 5 o o 0) 3 (0 E o !2 ^ 2 o o 55 1 Median _ "■- Upper 5% "^'^v^^^ ""--... Lower 5% 1963-70 73-74 78-79 83-84 88-89 93-94 1998-99 Fishing year Figure 6 The median and 90^"? confidence intervals of stock biomass predicted with the bootstrapped LMSE method for the period of 1969-70 to 1995-96. 0.30 0,25 - 0.20 - ^ Probab p p O en 1 , 0.05 -all lll...a 0 0.1 0.2 0.3 0.4 0.5 0.6 Ratio of 6,99=,^ to K Figure 7 The distribution of the ratio between the biomass of the stock in 1995-96 (Bjggs.gg) and the biomass of the virgin stock {K) estimated with the bootstrapped LMSE method. results for these two sets of data were reported in Mont- gomery and Chen ( 1996). We plotted the stock biomass estimated with these three sets of data (Fig. 10). Al- though there are differences, they are rather small con- sidering the large differences in adjustment coefficients. This may indicate that the results presented in this paper tend not to be sensitive to small changes in catch adjustment coefficients used in period II. 34 Fishery Bulletin 97(1), 1999 It is clear that the size of the lobster stock off New South Wales was low, with respect to its virgin bio- mass, as was the 75% confidence. Recent stock bio- mass (i.e. in 1995-96) has been between 15% and slightly over 30% of the virgin biomass. However, it seems that the decline in the biomass of the stock has stopped in recent years, and that the stock is perhaps in a period of recovery. 0.30 0.25 0.20 2 0.15 - o 0.10 0.05 0.00' ;,oco 0 500 1,000 1.500 2,000 2,500 Stock biomass (t) Figure 8 The distribution of stock biomass in 1996-97 estimated with the bootstrapped LMSE method. 1.0 ^_^^^=^ 'MSY -.^.i HI z> a. o 3 Q. O ally ignored these observations and tended to follow closely the CPUE in the majority of years (Fig. 11). Because of the high likelihood of under-estimation of CPUEs ft-om 1916 through 1924, the LMSE method is probably more suitable. The estimates of rjj, g^, and Sjggg_^o by the ML method were 0.17, 0.13 x 10"^ (per trap-month) and 7174 t, differing from those estimated with the LMSE and bootstrapped LMSE methods (Table 3). This difference may result from different weightings of data for years 1971-72 and 1974-75. Fitting the model with the LMSE method virtually ignored the data observed in these years, which had much higher CPUEs than other years. How- ever, fitting with the ML method was heavily influenced by these two years of high CPUEs (Fig. 11). Because of the patchy distribution of lobsters in their habitat and the expansion of fishing grounds out to the conti- nental slope off the NSW coast dur- ing the early 1970s, it is very likely that these two years of exception- ally high CPUEs resulted from high fishing efficiency (i.e. high (7 values), which should not be taken as an in- dicator of high biomass. We suggest using the bootstrapped LMSE method as an alternative approach to fitting production mod- els to catch-effort data. The results derived from such an analysis should be evaluated carefully with respect to the biology and ecology of the targeted fish species and with respect to how the catch-effort data were collected. Such an evaluation may shed some light on why some obsei-vations differ from the major- ity which the LMSE estimated line tends to follow. A comparison of re- sults between robust and traditional least squares approaches may lead to a better understanding of the dy- namics of the studied fish stock and identification of years in which atypical data are observed. 1940 1969-70 72-73 84-85 87-88 Fishing year Figure 11 The predicted catch per unit of effort with the maximum likehhood = 0.575j:-77.111 [r^=0.871] ^ 3 120 § 100 1 80 0 <^ ^ S 60- 40 20 ^^^^^'* n - ^n^njcp^^^^ u ^ 1 1 ' 1 1 100 150 200 250 300 350 Male length (iran) 180. 160- y = OAlSx - 50.380 [;^=0.839] 140- S 120- 1 100- <^ ^ Ji 80- 1 60- .S^--^^^ 40 ^^J^^'^ 20 ^ -,J;J»W» n * v^BO'ii^^^'^^^ 100 150 200 250 300 350 Farale length (nni) Figure 5 Regression of male length (mm) vs. male weight (g) (n = 119) and female length (mm) vs. female weight (g) (;i=96) for E. cimbrius collected during the GM 93-20 survey. laticauda and G. phycidis ranged from 1.26 to 1.34 mm and from 1.38 to 1.76 mm, respectively. Both digenetic trematodes were found in low prevalence (4.7% and 1.4%, respectively) (Table 6). The prevalence of infection of Raphidascarinidae sp. steadily increased from 8.3% in 1-year-olds to 100.0% in 9-year-olds (Table 7). The prevalence of G. erinaceus fluctuated but generally increased with fish age. There was an increase in infection from 9.1% to 20.0% and 25.0%, respectively by Raphidascarinidae sp. and G. erinaceus in 2- and 3-year-old fish. Infec- tion by G. laticauda and G. phycidis was minimal in all size classes. All helminths in all year classes were found in low abundance. Discussion Age and growth The age distribution of the sample comprised pri- marily of 3-, 4-, and 5-year-old E. cimbrius, with a maximum age of 9. The maximum age attained by this species in the Gulf of Maine is similar to that 46 Fishery Bulletin 97(1), 1999 180- 170- 160- ^ 150- * 140- 130 Age 4f ^ 120- „ * 9 110H X^,* 1 ^ 100- x_^^ ■ 8 Weight o o +x ^ , _ 70- + V X ^^t^ ,^+ O 5 60- 50 40- cw^ ^+ 30- ^^P [^- ° 3 20 1 r^-«a^^^ <^ 2 10H aur^"^^^*"- ' n| ^tH^^ , * 1 75 100 125 150 175 200 225 250 275 300 325 350 Total length (mm) Figure 6 Scattergram of age vs. total length (mm) and weight Ig) for£. cimbnus collected during the GM 93-20 survey. (n=241). reported from the western and eastern Atlantic (Cohen et al., 1990). Age classes discerned in this study also agree well with those documented by Cohen et al. (1990) for the eastern Atlantic. Age classes for the eastern Atlantic are defined as fol- lows: 3 years at 150 mm total length (TL), 5 years at 200 mm TL, 7 years at 250 mm TL, and 9 years at 290 mm TL (Cohen et al., 1990). The maximum length of E. cimbrius from this study (328 mm TL) again is simi- lar to the maximum length recorded in the western Atlantic (305 mm TL) by Cohen et al. (1990). The Kolmogorov-Smirnoff test showed no signifi- cant difference between length-frequency distribu- tions of observed total lengths and back-calculated lengths derived from otolith increment data. This finding further corroborates that presumed annuli in sagittal otoliths were formed at regular time in- tervals and can be used to assign ages to length classes. In aging studies, it is a common technique to com- pare assigned ages to length-frequency modes to con- firm age classes (Campana and Jones, 1992). This was not attempted because of substantial overlap in the length-frequency modes of E. cimbrius. Length- frequency modes with substantial overlap have proven to be unreliable as predictors of component age groups because of several factors: the number of modes may be less than the number of components, modes may be obscured or indistinct, or false modes may be generated (Beamish and McFarlane, 1987). Fourbeard recklings spawn from May to October (Battle. 1930; Cohen et al., 1990) and overlap m the length-fi-equency modes is likely due to this prolonged spawning season. After the first year, the age-length relationship of fourbeard recklings was linear. The young grew to an average of 113 mm during the first year, and, on average, grew 24 mm per year to age 9. The relative large size and narrow size range of one-year-old fish may be attributed to the under-representation of smaller fish in the sample size. Fast early growth, like that ofE. cimbrius, is exhibited by most gadids, and is particularly evident in red hake, Urophycis chuss, and white hake, U. tenuis. This type of growth is probably advantageous to U. chuss and U. tenuis and to other small demersal marine fishes that en- ter habitats dominated by larger individuals (Markle et al., 1982). Arctic cod, Boreogadus saida. from La- brador also exhibits growth rates similar to that of £. cimbrius. AHer reaching approximately 90 mm in their first year, B. saida average roughly 40 mm per year until age 5; they rarely live beyond age 6 (Lear, 1983). Deree; Age and growth, dietary habits, and parasitism of Enchelyopus cimbrius 47 Table 4 Relative importance of prey items oiE. cimbrius according to numerical and frequency of occurrence indices. (n=36). Prey taxon No. %No. Frequency %Frequency Phylum Mollusca Class Bivalvia 183 49.1 26 72.2 Order Protobranchia 183 49.1 26 72.2 Family Nuculanidae Yoldia sp. 183 49.1 26 72.2 Phylum Annehda Class Polychaeta 6 1.6 6 16.7 Family Phyllodocidae Unidentified Phyllodocidae 1 0.3 1 2.8 Paranaitis sp. 2 0.5 2 5.6 Family Nereidae Nereis sp. 1 0.3 1 2.8 Unidentified Polychaeta 2 0.5 2 5.6 Phylum Arthropoda Subphylum Mandibulata Class Crustacea 172 46.1 36 100.0 Order Copepoda 138 36.9 8 22.2 Suborder Calanoida 138 36.9 8 22.2 Subclass Malacostraca Series Eumalacostraca Superorder Peracarida Order Cumacea 13 3.5 9 25.0 Eudorella truncata 1 0.3 1 2.8 Diastylis quadrispinosa 7 1.9 4 11.1 Unidentified Cumacea 5 1.3 4 11.1 Order Isopoda 2 0.5 2 5.6 Unidentified Isopoda 2 0.5 2 14.0 Order Amphipoda 5 1.6 5 13.9 Family Hyperiidea Unidentified Hyperiidea 3 0.8 2 5.6 Family Calliopidae Calliopius laeviusculus 1 0.3 1 2.8 Family Gammaridae Unidentified Gammaridae 2 0.5 2 5.6 Order Mysidacea 1 0.3 1 2.8 Mysidopsis bigelowi 1 0.3 1 2.8 Order Decapoda 12 3.2 11 30.6 Infraorder Caridea Family Crangonidae Crangon septemspmosa 5 1.3 4 11.1 Family Pandalidae Unidentified Pandalidae 1 0.3 1 2.8 Unidentified Decapoda 4 1.1 4 11.1 Infraorder Brachyura Unidentified Brachyura 2 0.5 2 5.6 Animal remains 11 2.9 10 27.8 Unidentified vertebrates 1 0.3 1 2.8 Number examined 36 Number empty 3 Total length is the most appropriate parameter to set limits around an age class for E. cimbrius. For example, a fish with a total length of 165 mm would be approximately 3 or 4 years old. If the weight of this same fish, approximately 15 g, was used to esti- mate age, the range of this estimated age would in- crease from 2 to 5 years. Conversely, weight appears to be a better measure of absolute growth than total 48 Fishery Bulletin 97(1), 1999 Table 5 Relative importance of prey items of 1 -year-old 1 105- 121 mm TL i;!=5i and 2- -7 year old (119- -271 mm TLl {n=3l}E. cimbniifi based on numerical and freq uency of occurrence indices. Prey taxon 105-121 mmTL 119- 271 mm TL No. %No. Freq. 'J'fFreq. No. %No. Freq. %Freq. Phylum Mollusca Class Bivalvia 14 19.2 4 80.0 169 56,3 22 71.0 Order Protobranchia 14 19.2 4 80.0 169 56.3 22 71.0 Family Nuculanidae Yoldia sp. 14 19.2 4 80.0 169 56.3 22 71.0 Phylum Annelida Class Polychaeta 0 0.0 0 0.0 6 2.0 6 19.4 Family Phyllodocidae Unidentified Phyllodocidae 0 0 1 0.3 1 3.2 Paranaitia sp. 0 0 2 0.7 2 6.5 Family Nereidae Nereis sp. 0 0 1 0.3 1 3.2 Unidentified Polychaeta 0 0 2 0.7 2 6.5 Phylum Arthropoda Subphylum Mandibulata Class Crustacea 59 80.7 8 160.0 113 37.7 28 90.3 Order Copepoda 52 71.2 4 80.0 86 28.7 4 12.9 Suborder Calanoida 52 71.2 4 80.0 86 28.7 4 12.9 Subclass Malacostraca Series Eumalacostraca Superorder Peracarida Order Cumacea 4 5.4 2 40.0 9 3.0 7 22.6 Eudorelta truncata 0 0 1 0.3 1 3.2 Diastylis qua-« w X X x-x ♦ » 0 /i M M M I M I M M M I! I I X I I I I M M I M I M I M iV-X-X-X-X-X 95 115 135 155 175 195 215 235 255 275 295 315 Total length (mm) Figure 7 Length-frequency distribution of back-calculated lengths at age vs. observed lengths at age of £. cimbrius collected during the GM 93-20 survey between 2 August and 13 August 1993 plotted in groups of 5. The dietary habits of .E. cimbrius may contribute to its constant absolute growth. Fourbeard rocklings generally forage on small animals in large numbers instead of large animals in small numbers. Fishes that swim actively and forage for small organisms (i.e. "grazers") generally have a higher unit of en- ergy expenditure per unit of energy input than do fishes that are lie-and-wait predators on large or- ganisms. In addition, preference for cold tempera- tures may explain further why E. cimbrius exhibits constant linear growth. I speculate that£. cimbrius uses its energy to forage; energy that would other- wise have contributed to absolute growth. Table 6 Prevalence (P), abundance (A), standard error (SE). and intensity (I) of Raphidascaridinae sp., Grillotia erinaceus. Genolmea laticauda, and Gonocerca phycidis in E. cimbrius collected during the GM 93-20 and GM 94-12 surveys. Parasite P(%) A + SE I range Raphidascaridinae sp. Grillotia erinaceus Genolinea laticauda Gonocerca phycidis 4 23.0 1.69 ± 1.24 1-32 1 20.9 1.65 ± 0.25 1-8 7 4.7 4.33 ± 0.96 1-7 1 1.4 2.00 ± 0.00 2 Dietary habits Prey items ofE. cimbrius surveyed in this study were predominately bivalves, particularly Yoldia sp., in both number and frequency. Calanoid copepods, although present in great numbers, were consumed by few indi- viduals. The importance of copepods is replaced by bivalves, Yoldia sp., and polychaetes in the diet of larger fish ( > 1 19 mm TL ). After their first year, fish shift from a diet dominated by planktonic copepods to a diet com- prising infauna (bivalves and polychaetes). Calanoid copepods composed the majority (71.2%) of the prey taken, with bivalves (although consumed in relatively high frequency (80.0%)) second in im- portance by number ( 19.2%) in 1-year-old fish. Juve- niles in the present study exhibited feeding prefer- ences for various calanoid copepods similar to those of pelagic larval gadids off the Isle of Man (Nagab- hushanam, 1965.) This preference for planktonic or- ganisms suggests that juveniles lead a less seden- tary existence and feed more frequently in the water column than do adults. In addition, the selection for smaller prey is directly related to mouth size (Moyle and Cech, 1988). On Georges Bank, fourbeard rocklings >100 mm TL consumed motile epifauna, particularly Crangon septemspinosa (46.3%), in the highest percentage by 50 Fishery Bulletin 97(1), 1999 Table 7 Prevalence of infection (P%) and abundance (A) of four species of helminths in age-c ass ranges Tom E. cimbrius collected during | the GM 93-20 and GM 94-12 surveys. (n=148). Age class n Raphidascaridinae sp. G. erinaceus G. laticauda G. phycidis P(%) A P(%) A P(%) A P(%) A 1 12 8.3 0.1 16.7 0.2 0.0 0.0 0.0 0.0 2 11 9.1 0.5 9.1 0.2 9.1 0.2 9.1 0.2 3 20 20.0 0.3 25.0 0.3 0.0 0.0 0.0 0.0 4 51 21.6 1.3 17.6 0.2 5.9 0.2 0.0 0.0 5 27 26.0 2.3 14.8 0.4 3.7 0.2 0.0 0.0 6 15 26.7 1.9 20.0 0.4 6.7 0.5 0.0 0.0 7 8 37.5 1.5 37.5 0.4 0.0 0.0 0.0 0.0 8 2 50.0 1.0 100.0 1.5 0.0 0.0 0.0 0.0 9 2 100.0 2.5 100.0 2.5 0.0 0.0 0.0 0.0 weight followed by animal remains (30.1'^) and in- fauna, primarily polychaetes (13.6%) (Langton and Bowman, 1980). Decapod crustaceans were not as im- portant as these other groups to juvenile fish in the present study. Decapods appeared in number for the first time in 2- to 7-year-old fish. This can be explained by the preference for larger prey by older, larger fish. In a Norwegian Qord, E. cimbrius between 70 and 300 mm TL preyed primarily on crustaceans, poly- chaetes, fish, bivalves and rarely on copepods (Mattson, 1981). In adult rocklings, high percentages of infauna, particularly polychaetes, have been re- ported from the North Sea (Moller-Buchner et al., 1984) and Passamaquoddy Bay, where the frequency was found to be correlated with an increase in fish size (Keats and Steel, 1990). The high frequency of infaunal organisms, predominantly bivalves, in the diet of 2- to 7-year-old fish in the present study ex- hibits a similar trend. This finding may be correlated with adults having a sedentary existence and remain- ing in close contact with the substrate (Bigelow and Schroeder, 1953; Cohen et al., 1990). This lifestyle, compounded with the fact that this species has well- developed stout barbels equipped with taste buds (Nagabhushanam, 1965), would enable larger fish to feed more effectively on epifauna and infauna. E. cimbrius is best characterized as a euryphagous bottom-rover or "grazer." (Moyle and Cech, 1988). The size of the prey appears to be directly correlated with the size of the predator as seen in the change in diet from 1-year-old fish to those 2 to 7 years of age. Parasitism The parasite fauna of £. cimbrius consisted of three taxa and four species: Nematoda ( 1 ), Cestoda ( 1 ), and Trematoda (2). The prevalence (P^ ) and abundance (A) of a nematode from the subfamily Raphidas- caridinae (probably Hysterothylacium) and a trypanorhynch cestode, G. erinaceus, appear to in- crease with fish age. The P% and A of both digenetic trematodes were low in all age classes. The few helminths found were in low abundance. This is uncommon because other gadids are known to have abundant and diverse parasite faunas (Margolis and Arthur, 1979). The very low parasite diversity of E. cimbrius supports evidence of limited movements and benthic feeding preferences in its habitat. All nematode specimens of the subfamily Raph- idascaridinae were second- or third-stage and third- or fourth-stage larvae (Measures^. The presence of parasites in the same stage of development are in- dicative of a "grazer" (Dogiel, 1966) and provides additional support to characterize E. cimbrius as such. A similar trend is seen in G. morhua, where its mode of life, generalist feeding habits, and associa- tion with the bottom facilitate the infestation by Raphidascaridinae spp. (Dogiel et al., 1970). The increase in prevalence of Grillotia erinaceus from 9.1% at 2 years to 25.0% at 3 years appears to be correlated with the an increase of arthropods in the diet. In the life cycle of G. erinaceus, the plerocercus larvae can be acquired by an intermedi- ate teleost host through consumption of arthropods, mainly copepods, which have previously ingested the tapeworm eggs (Sakanari and Moser, 1985). It is not possible to draw any firm conclusions on this rela- tionship because food habit data in this study did ' Measures, L. 1995. Maurice Lamontagne Institute, Fisher- ies and Oceans, Mont-Joli (Quebec), Canada G5H 3Z4. Per- sonal comniun. Deree: Age and growth, dietary habits, and parasitism of Enchelyopus cimbhus 51 not account for seasonal variation and because the sample size of 8- and 9-year-old fish was low. The definitive hosts of adult trypanorhynchs are elasmobranchs, suggesting that fourbeard recklings are consumed by elasmobranchs or act as paratenic hosts. It is likely that barndoor skate. Raja laevis, and thorny skate, Raja radiata, prey upon E. cimbrius. Raja laevis and R. radiata >700 mm TL forage primarily on fish that include hake, Urophycis spp., and whiting, Merluccius spp., and may also feed on cod and haddock (Bigelow and Schroeder, 1953; McEachran et al., 1976). Because both jRqya spp. con- sume fishes with similar habits and are definitive hosts for G. erinaceus (Margolis and Arthur, 1979), it is prob- able that/?, laevis andi?. radiata prey upon E. cimbrius. Acknowledgments I am indebted to. J. G. Hoff, S. A. Moss, R. A. Campbell, and K. Oliveira for their critical review and guidance throughout this project. I am grateful to the National Marine Fisheries Service Survey and the Age and Growth Divisions in Woods Hole, MA, and to the Massachusetts Division of Marine Fish- eries for their collection assistance and use of equip- ment. I would like to thank L. Measures for her assis- tance in nematode identification. Special thanks are given to L. Curran, R. P. Glenn, and J. A. and J. Fowler for their efforts. This manuscript was part of a thesis submitted in partial fulfillment of the requirements for the M.S. degree in marine biology at the University of Massachusetts, Dartmouth, and is dedicated to M. A. Rousseau and V. A. Nordahl for inspiration. Literature cited Andrade, K. G., and C. P. Smith. 1988. Pollock, Pollachius virens. In J. Penttila and L. M. Dery (eds.). Age determination methods for Northwest At- lantic species, p. 37-40. U.S. Dep. Commer, NOAA Tech Rep NMFS 72. Battle, H. I. 1929. Effects of extreme temperatures and salinities on the development ofEnchelyopus cimbrius (L.). Contrib. Can. Biol. Fish., new series, V(6l:107-192. 1930. Spawning periodicity and embryonic death rate of £. cimbrius (L.) in Passamaquoddy Bay. Contrib. Can. Biol. Fish., new series, V;363-380. (also V(6):13.; V(7):361-380.) Beamish, R. J., and G. A. McFarlane. 1987. Current trends in age determination methodol- ogy. In R. C. Summerfelt and G. E. Hall, (eds.), The age and growth offish, p. 15-42. The Iowa State Univ. Press, Ames, lA. Bigelow, H. B., and W. C. Schroeder. 1953. Fishes of the Gulf of Maine. U. S. Fish. Wildl. Serv. Fish. Bull. 74:234-238. Brinkmann, A., Jr. 1988. Presence of Otodistomum sp. metacercariae in Nor- wegian marine fishes. Sarsia. 73:79-82. Campana, S. E., and C. M. Jones. 1992. Analysis of otolith microstructure data. In D. K. Stevenson and S. E. Campana (eds. I Otolith microstruc- ture examination and analysis, p. 73-100. Can. Spec. Publ. Fish. Aquat. Sci. 117. Carlander, K. D. 1981. Caution of the use of the regression method of back- calculating lengths from scale measurements. Fisheries. 6:2-4. Cohen, D. M., T. Inada, T. Iwamoto, and N. Scialabba. 1990. Gadiform fishes ofthe world ( Order Gadiformes). FAO Fish. S>Tiop. 125. vol. 10:38-39. FAO, Rome. Demir, N., A. J. Southward and P. R. Dando. 1985. Comparative notes on postlarvae and pelagic juve- niles ofthe recklings Gaidropsaurus mediterraneus, Rhino- nemus cimbrius. Ciliata mustela and C. septentrionalis. J. Mar. Biol. Assoc. U. K. 65(31:801-839. Dery, L. M. 1988. Silver hake, Merluccius bilinearis. In J. Penttila and L. M. Dery (eds.). Age determination methods for Northwest Atlantic species, p. 41-48. U.S. Dep. Commer, NOAA Tech Rep NMFS 72. Dogiel, V. A. 1966. General parasitology, revised and enlarged by Y. I. Polyanski and E. M. Kheisin. Academic Press. New York, NY. 516 p. Dogiel, V. A, G. K. Pertushevski, and Y. I. Polyanski (eds.). 1970. Parasitology of fishes. T. H. F Publications, Inc. Ltd., Neptune City, NJ, 384 p. Fournier, D. A. 1983. Analysis of otolith microstructure data. In D. K. Stevenson and S. E. Campana (eds.). Otolith microstruc- ture examination and analysis, p. 73-100. Can. Spec. Publ. Fish. Aquat. Sci. 117. Gibson, D. I., and R. A. Bray. 1984. On Anomalotrema Zhukov, 1957, Pellamyzon Mont- gomery. 1957. and Opecoelina Manter, 1934 (Digenea: Opecoelidae). with a description of Anomalotrema koiae sp. nov from North Atlantic waters. J. Nat. Hist. 18:949-964. Keats, D. W., and D. H. Steele. 1990. The fourbeard rockling, Enchelyopus cimbrius (L.), in eastern Newfoundland. J. Fish. Biol. 37(5):803-811. Langton, R. W., and R. E. Bowman. 1980. Food of fifteen northwest Atlantic gadiform fishes. U.S. Dep. Commer., NOAA Tech. Rep. NMFS. SSRF 740. 23 p. Lear, W. H. 1983. Arctic cod: underwater world. In W. B. Scott and M. G. Scott (eds.), Atlantic fishes of Canada, p. 260- 262. Can. Fish. Bull. Aquat. Sci. 219. Linton, E. 1899. Fish parasites collected at Woods Hole in 1898. Bull. Bur. Fish. Vol. XIX:267-304. Lorn, J., and M. Laird. 1969. Parasitic protozoa from marine and euryhaline fish of Newfoundland and New Brunswick. I. Peritrichous ciliates. Can. J. Zool. 47:1367-1380. Margolis, L., and J. R. Arthur. 1979. Synopsis of the parasites of the fishes of Canada. Bull. Fish. Res. Board Can. 199, 269 p. MargoUs, L., G. W. Esch, J. C. Holmes, A, M. Kuris, G. A. Shad. 1982. The use of ecological terms in parasitology. (Report 52 Fishery Bulletin 97(1), 1999 of an AD HOC committee of the American Society of Parasitologists). J. Parasitol. 685(1):131-133. Markle, D. F., D. A. Methven, and L. J. Coates-Markle. 1982. Aspects of spatial and temporal cooccurrence in the life history stages of the sibling hakes, Urophycis chuss (Walbaum 1792) and Urophycis tenuis (Mitchill 1815) (Pisces:Gadidae). Can. J. Zool. 60:2057-2078. Mattson, S. 1981. The food o(Galeus melastomus. Gadiculus argenteus thon, Trisopterus esmarkii, Rhinonemiis cimbnus, and Glyptocephalus cynoglossus (Pisces) caught during the day with shrimp trawl in a West-Norwegian fjord. Sarsia. 66:109-127, McEachran, J. D., D. F. Boesch, and J. A. Musick. 1976. Food division within two sympatric species-pairs of skates (Pisces:Rajidae). Mar. Biol. 35:301-317. MoIIcr-Buchner, J., C .D. Zander, and D. Westphal. 1984. On the feeding habits of some demersal and suprademersal fish from Fladen Ground, North Sea. Zool Anz. 213(1-2):128-144. Moyle, P. B., and J. J. Cech Jr. 1988. Fishes: an introduction to ichthyology. Prentice Hall, Englewood Cliffs, NJ, 559 p. Nagabhushanam, A. K. 1965. On the biology of the commoner gadoids in Manx waters. J. Mar Biol. Assoc. U. K. 45:615-657. Odense, P. H., and V. H. Logan. 1976. Prevalence and morphology o{ Eimeria gadi (Fiebiger 1913) in the haddock. J. Protozool. 23(4):564-571. Pentilla, J. 1988. Atlantic cod, Gadus morhua. In J. Penttila and L. M. Dery (eds.), Age determination methods for Northwest Atlantic species, p. 31-36. U.S. Dep. Commer., NOAA Tech Rep NMFS 72. Pentilla, J., F. Nichy, J. Ropes, L. Dery, and A. Jerald Jr. 1988. Methods and equipment. In J. Penttila and L. M. Dery (eds.). Age determination methods for Northwest At- lantic species, p. 7-16. U.S. Dep. Commer, NOAA Tech Rep NMFS 72. Sakanari, J., and M. Moser. 1985. Infectivity of, and laboratory infection with, an elas- mobranch cestode, Lacistorhynchus tenius (Van Beneden, 1858). J. Parasitol. 71 (6):788-791. Secor, D. H., J. M. Dean, and E. H. Laban. 1992. Otolith removal and preparation for microstructural examination. In D. K. Stevenson and S. E. Campana (eds.). Otolith microstructure examination and analysis, p. 19-57. Can. Spec. Publ. Fish. Aquat. Sci. Tully, C, and P. O Ceidigh. 1989. The ichthyoneuston of Galway Bay ( west of Ireland ). II. Food of post-larval and juvenile neustonic and pseudoneustonic fish. Mar. Ecol. Prog. Ser. 51(3);301-310. Tyler, A. V. 1972. Food resources division among northern, marine, demersal fishes. J. Fish. Res. Board Can. 29:997-1003. Wheeler, A. 1969. The fishes of the British Isles and North-west Europe. Michigan State Univ. Press, East Lansing, MI, 613 p. 53 Abstract.— Nucleotide sequences of a 394-396 base pair fragment of mito- chondrial (mt) DNA, including parts of the cytochrome b and threonine tRNA genes, were obtained for eleven species of carcharhiniform sharks important to the U.S. Atlantic large coastal shark fishery. Sequences were used to predict sizes of restriction fragments produced by 118 restriction enzymes with unique recognition sequences. Seven restric- tion enzymes were chosen that produce an array of species-specific fragments for the eleven species. Geographic variation was examined in several spe- cies by surveying specimens from geo- graphically distant regions. Only one of the species, the spinner shark (Carcharhmus brevipinna), exhibited geographic variation in mtDNA restric- tion fragments. The sandbar shark (C. p/(/m6eus) exhibited sequence polymor- phism that did not produce differences in restriction patterns of any of the seven enzymes. We detected numerous differences between observed restric- tion patterns in ten tiger sharks (Galeocerdo cuvier) and patterns pre- dicted from a published sequence. We concluded that the published sequence is incorrect. Amplification of a single PCR product from a sample of meat, digestion of aliquots of the product with restriction enzymes, and sizing of frag- ments on agarose gels is an efficient method for distinguishing among these eleven carcharhiniform sharks. The method can be applied when only a small amount of tissue is available. Genetic identification of sharks in the U.S. Atlantic large coastal shark fishery* Edward J. Heist John R. Gold Center for Biosystematics and Biodiversity Texas A&M University, College Station, Texas 77843-2258 Present address: Cooperative Fisheries Research Laboratory Department of Zoology Southern Illinois University Carbondale, Illinois 62901-6511 E-mail addresss (for E J Heist): edheist@siu.edu Manuscript accepted 19 March 1998. Fish, Bull. 97:53-61 (1999). The U.S. Atlantic large coastal shark fishery grew rapidly during the 1980s when commercial landings in- creased from 135 metric tons (t) ih 1979 to a high of 7122 t in 1989 (NMFS, 1993). In 1993, a quota of 2750 1 was established; in 1997 this quota was halved to 1375 t in order to rebuild depleted shark stocks. This fishery targets several species of sharks that are valued for fins (exported to Asia) and meat (sold domestically). Because of the slow growth rate, high age at maturity, and low fecundity of most shark species (Pratt and Casey, 1990), commercial shark fisheries typically collapse after a brief period unless strict conservation measures are implemented (Holden, 1974, 1977; Hoenig and Gruber, 1990). The 1993 shark fishery management plan di- vided exploited species into three categories: large coastal sharks, small coastal sharks, and pelagic sharks (NMFS, 1993). The category that grew most rapidly was the large coastal shark fishery, which is dominated by several species of requiem sharks and two species of hammerhead sharks. Sound management of a multi- species fishery requires information on the vulnerability of each compo- nent of the fishery. Differences in life history characters, e.g. intrin- sic growth rates, locations of nurs- ery areas, or migration within or outside of the fished area, can re- sult in different vulnerabilities to overfishing of exploited species. For example, Musick et al. (1993) re- ported a relative decline in the dusky shark (Carcharhinus obscurus) off Virginia during expansion of the large coastal shark fishery. Thus, it is important to estimate catches on a species-by-species basis and to implement species-specific manage- ment. In the event that regulations (e.g. moratoria or minimum size limits) are applied to individual spe- cies, enforcement will rely on iden- tification of protected species within the catch. The manner in which sharks are processed at sea, how- ever, makes it difficult to accurately identify species at landing. Sharks typically are headed, gutted, and finned (i.e. fins are removed), thus destroying morphological charac- ters necessary for species identifi- cation. Although Castro ( 1993) rec- ommended a suite of characters for identification of shark carcasses, the limited number of available morphological characters makes it difficult to distinguish among sev- eral species. Martin (1993) sug- gested that use of restriction-frag- * This paper represents number XVIII in the series "Genetic studies in marine fishes" and contribution number 48 of the Center for Biosystematics and Biodiversity at Texas A&M University, College Station, Texas 77843-2258. 54 Fishery Bulletin 97(1), 1999 ment differences in polymerase chain reaction (PCR) amplified DNA might provide a rapid and inexpen- sive means of identifying carcasses and fins. We de- velop this technique as a means of identifying the eleven most frequently landed carcharhiniform sharks in the U.S. east coast longline fishery. Materials and methods Tissues (heart, white muscle, or fin ) from nine spe- cies of carcharhinid sharks and two species of sphyrnid sharks were obtained from commercial fish- ermen, sport tournaments, and research longlining cruises (Table 1). Tissues were either frozen in the field and stored at -80°C or preserved immediately in lOx Longmire's lysis buffer (0.1 M tris, 0.1 M NagEDTA, 0.01 M NaCl, 0.5% SDS, pH 8.0) at room temperature. Genomic DNA was isolated by first powdering tissue under liquid nitrogen with a prechilled mortar and pestle. Approximately 100 mg of powdered tissue were suspended in 500 |aL STE buffer (0.1 M NaCl, 50 mM tris, 1 mM EDTA; pH 7.5) and lysed with 25 |aL 20% SDS. Genomic DNA was extracted twice with phenol:chloroform:isoamyl alcohol (25:24:1) and twice with chloroform:isoamyl alcohol (24:1). DNA in aqueous phase was precipi- tated by adding 2.5 volumes of ice-cold absolute etha- nol and 0.1 volumes of 3M NaOAC, stored at -20°C for at least two hours, centrifuged for 15 min at 4°C at maximum speed in a microcentrifuge, and rinsed with 70% ethanol. PCR amplification for cycle sequencing or restric- tion-enzyme digestion was accomplished by using a suite of PCR primers. The "diagnostic" fragment used in digestions consisted of a single segment, 394-396 bp in length, that was amplified by using light-strand primer Cb3RL (CATATTAAACCCGAATGATAYTT) located within the 3' domain of the mitochondrially encoded cytochrome b (cyt b) gene and heavy-strand primer Cb6H (CTCCAGTCTTCGRCTTACAAG) lo- cated within the mitochondrially encoded threonine tRNA (tRNA™!^) gene (Martin and Palumbi, 1993). This fragment was sequenced through the primer sites by using additional primer sets within and out- side the diagnostic fragment. PCR fragments were prepared for sequencing by using the Bio-Rad Prep- a-Gene DNA purification system that removes ex- traneous salts, primers, and small fragments of DNA prior to cycle sequencing. Dideoxy DNA sequencing was performed with the Promega fmol DNA sequenc- ing system by using ■^■^P end-labeled primers. Cycle sequencing reactions consisted of a single two-minute denaturing process at 95°C, followed by thirty cycles of 1 min at 95=C, 30 sec at 64°C, and 30 sec at 72°C. Sequencing reactions were scored on 6% denaturing Table 1 Sources of tissues or DNA sequences. EMBL/Genbank accession numbers identify sequences from Martin and Palumbi (1993). Species Acronym Source of sequence Source of specimen Bignose shark (Carcharhinus altimus) Cal-A This study Virginia, Atlantic Ocean Blacktip shark (C. limbatus) Cli-A " Virginia, Atlantic Ocean Bull shark (C. leucas) Cle-A " Florida, Gulf of Mexico Dusky shark (C. obscurus) Cob-A " Virginia, Atlantic Ocean Sandbar shark (Al (C. plumheus) Cpl-A GenBank L08032 Hawaii. Pacific Ocean Sandbar shark (B) Cpl-B This study Florida. Gulf of Mexico Sandbar shark (C) Cpl-C " Hawaii. Pacific Ocean Silky shark (C falciformis) Cfa-A II Hawaii. Pacific Ocean Spinner shark (A) (C. brevipinna) Cbr-A " Pacific Ocean, Australia Spinner shark (B) Cbr-B " Virginia, Atlantic Ocean Lemon shark iNegaprion brevirostris) Nbr-B GenBank L08039 Florida, Atlantic Ocean "Tiger shark" (A)' (Galeocerdo cuvier) Gcu-A GenBank L08034 Hawaii, Pacific Ocean Tiger shark (B) Gcu-A This study GenBank AF004288 Florida. Gulf of Mexico Great hammerhead (Sphyrna moka'rran I Smo-A '■ Florida. Gulf of Mexico Scalloped hammerhead (S. lewini) Sle-A GenBank L08041 Hawaii, Pacific Ocean ' It was determined that this sequence was not tiger shark 'see text' Heist and Gold: Genetic identification of sharks 55 polyacrylamide gels and examined by autoradiogra- phy. DNA sequences were read on an IBI gel reader and entered directly into computer text files. Se- quences were confirmed by recording a minimum of two separate sequencing runs through each base. Sequences were aligned by using the ESEE software package (Cabot and Beckenback, 1989). Four cyt b sequences — those of sandbar (C. plumbeus), lemon (Negaprion brevirostris), tiger (Galeocerdo cuvier), and great hammerhead sharks {Sphyrna mokarran ) — were downloaded from the EMBL/NCBI Genbank database (Martin and Palumbi, 1993). All other cyt b sequences and all tRNA^""^ sequences were acquired from our laboratory. Sequences from each species between primers Cb3RL and Cb6H were concatenated with sequences of the primers to produce a single sequence for each of the eleven species (Table 2). Sequences were exam- ined for predicted restriction sites with IBI MacVector software (IBI Mac Vector, 1991). One hundred eigh- teen restriction enzymes with unique recognition se- quences were used in the search for restriction sites. Amplified "diagnostic" fragments were prepared by amplifying genomic DNA at thirty cycles of 95°C for 1 min, 48 to 52°C for 30 sec, and 72°C for 30 sec. Amplified fragments were ethanol precipitated as above and reconstituted in water. Generally, one 100 |iL PCR reaction produced enough diagnostic frag- ment for all seven restriction digestions. Fragments were digested by using manufacturer's buffers and specifications. Most restriction patterns were scored on 2% agarose gels run on IX TAE. Fragments be- tween 40 and 100 bp were scored on either vertical nondenaturing polyacrylamide gels or 2'7( nusieve 3:1 agarose gels. Fragments less that 40 bp were not scored, although loss of fragments as small as 24 bp could be inferred from mobility shifts in larger frag- ments. All gels were stained with ethidium bromide and examined under UV light (Sambrook et al., 1989). Results A single mtDNA fragment of 394-396 bp that con- tained the 3' end of cyt b and part of the tRNATHR gene was amplified in all species. Except for two ham- merhead sharks, the fragment was 395 bp in length (Table 2). Cyt b sequences from all eleven species were identical in length and unambiguously aligned. In comparison with the other nine species, the scal- loped hammerhead (S. lewini) possessed a single- base deletion in the tRNA^"'^ sequence, and the great hammerhead possessed a single-base insertion in the tRNA''""'^ sequence. In bignose shark (C. altimus), three fragments of sizes 395 bp, 720 bp, and 1040 bp were produced re- peatedly. The additional bands were more pro- nounced at lower (48°C) than higher (52°C) anneal- ing temperatures. Further investigation with addi- tional primers revealed that bignose shark possessed a much larger mitochondrial D-loop region than any other shark in the study. Amplification with a light- strand primer located within cyt b and a heavy- strand primer located within the 12S ribosomal RNA gene produced an approximately 1400-bp fragment in all species except bignose shark, where a single fragment of approximately 2000 bp was produced. We hypothesize that the Cb6H recognition site within the tRNA^"^ gene is duplicated twice within the D- loop of the bignose shark, resulting in a three-banded amplification product. Duplication of segments of flanking regions within the mitochondrial D-loop have been reported in other vertebrates (Broughton and Dowling, 1994, and citations within). To our knowledge, this is the first evidence of this phenom- enon in elasmobranchs. To obtain the single product for sequencing and restricting, the 395-bp fragment was excised from an agarose gel and reamplified to produce sufficient amounts of the diagnostic fragment. Of 1 18 restriction enzymes surveyed by MacVector software, 34 were predicted to have restriction sites within one or more of the eleven species. Seven re- striction enzymes (Alul, Ddel. Fokl, Haelll, Hindi, Hinfl, and Rsal) were chosen for screening because use of these seven enzymes allowed all eleven spe- cies to be distinguished (Table 3). The number of sharks whose mtDNA was subjected to restriction- enzyme digestion is given in Table 4. The initial screen of restriction sites was undertaken with only a single sequence from each species. Differences in predicted and observed restriction patterns in three species (sandbar, spinner, and tiger sharks) made it necessary to sequence additional animals in order to investigate whether differences were due to sequenc- ing errors or intraspecific variation. Observed restriction patterns in sandbar sharks from the Gulf of Mexico differed by restriction site from the pattern predicted from the published se- quence of Martin and Palumbi (1993) for a sandbar shark (Cpl-A) from Hawaii (MartinM. Although we predicted that Fokl would not cut the sandbar shark fragment, Fokl digestion produced two fragments of 310 and 85 bp. We sequenced additional sandbar sharks from the Gulf of Mexico (Cpl-B) and from Hawaii (Cpl-C) and found that both sequences pos- sessed theFo^I restriction site. Sequence of the speci- ' Martin, A. P. 1997. University of Nevada-Las Vegas, Las Ve- gas, NV 89154-4004. Personal commun. 56 Fishery Bulletin 97(1), 1999 Table 2 DNA Sequences used to predict restriction digests. Boxes surround primer sequences used in PCR amplification (A=adenine, C=cytosine, G=guanine, T=thymidine, Y=C or T). A vertical line is drawn between the end of the cj^ochrome 6 sequence and the beginning of the threonine tRNA sequence. See Table 1 for description of samples. 5' Cb3RL -> Cal-A Cbr-A Cbr-B Cfa-A Cle-A Cli-A Cob-A Cpl-A Cpl-B Cpl-C Gcu-A Gcu-B Nbr-A Sle-A Smo-A Cal-A Cbr-A Cbr-B Cfa-A Cle-A Cli-A Cob-A Cpl-A Cpl-B Cpl-C Gcu-A Gcu-B Nbr-A Sle-A Smo-A CATATTAAACCCGAATGATAYTT CTTATTTGCTTATGCAATCCTGCGCTCAATCCCTAATAAACTAGGAGGAGTCCTAGC C T.A C C T.A C C C T.A T A C C..C A T TT . A C .C C T.C..G C T.A..T..T C T C A T..T C .C....C..C T.A T..T C T.... C..C T.A C..T C TCTCCTATTCTCTATCTTCATCCTTATATTGGTGCCCCTCCTCCACACCTCCAAACAACGAAGTACCATCTTCCGACCCA C T..T C A..C..T T T . . .T. .T '. .C . . .C. . .G . .C. C T TC . T C . . AG T C . C C..T C. .A. .C. . . . .T. .A. . T. . AA . T . .A. .T. . . .A. .T. . C. AA.T. •C.A. .C . . C .G. .C . C . . TT . • C.A. • C.A. .T..A..A..T A. . C . TTTA. ..T T A. . .A. . AAC .AC .AC continued men from the Gulf of Mexico (Cpl-B) differed from that of the specimen from Hawaii (Cpl-C) by three base-pair substitutions. The sequence from the speci- men from Hawaii (Cpl-C) differed from that of Martin and Palumbi (1993) by one base-pair (and which af- fected thefo^I restriction site). Whether the single base pair difference between our specimen from Hawaii and the one reported in Martin and Palumbi ( 1993) repre- sents an uncommon polymorphism in sandbar sharks from the Pacific or an error in reading the original se- quence cannot be determined. We examined restriction digestions from three sandbar sharks collected near Hawaii and all possessed the Fokl restriction site. We also found a populational difference between spinner sharks (C. brevipinna) collected from the North Atlantic (including the Gulf of Mexico) and from the Pacific coast of Australia. Initially, we se- quenced a spinner shark from Australia and pre- dicted that Rsal would not cut the diagnostic frag- ment. We restricted fragments from seven spinner sharks, two from Australia, and five from the U.S. Atlantic and Gulf of Mexico coasts. Although frag- ments from the two spinner sharks from Australia were not cut with Rsa I, each of the five specimens from the Atlantic produced fragments of 251 and 144 bp. We sequenced one specimen from the Atlantic (spin- ner B) and found two nucleotide substitutions, includ- ing one that resulted in the restriction site difference between spinner sharks from the Atlantic and Pacific. We detected numerous differences between restric- tion patterns predicted from the sequence of Martin and Palumbi ( 1993 ) for tiger shark collected near Ha- waii and our tiger sharks collected from the Gulf of Mexico (Atlantic) and Hawaii and Australia (Pacific). Heist and Gold: Genetic identification of sharks 57 Table 2 (continued) Cal-A Cbr-A Cbr-B Cfa-A Cle-A Cli-A Cob-A Cpl-A Cpl-B Cpl-C Gcu-A Gcu-B Nbr-A Sle-A Smo-A Cal-A Cbr-A Cbr-B Cfa-A Cle-A Cli-A Cob-A Cpl-A Cpl-B Cpl-C Gcu-A Gcu-B Nbr-A Sle-A Smo-A Cal-A Cbr-A Cbr-B Cfa-A Cle-A Cli-A Cob-A Cpl-A Cpl-B Cpl-C Gcu-A Gcu-B Nbr-A Sle-A Smo-A TAACACAAATCTT T . . CTTC rGACTTCTTGTGGCCAACTCAATTATTTTAACTTGAATTGGAGGTCAACCAGTAGAACAACCA A-.T-.T C T . A T . . T . . . . c . . c . . A , C . . . . . . . c . . r . A . . T . . T . . T . A . . T . . T . . . . . C . . C . A . . T . . . . . c . . C C . . . T . C . . A . . T . . . . C . . CC . . . . . . .c . . r . A . . T . C . . . . A. . . . . T . .A . . T . . . .c . . r . A . CC . . . . . .c . . T - r . A . . T . c. . . . r TTCATTATAGTAGGACAAATCGCCTCAATCTCCTACTTTTCCTTATTCCTTATTATTATACCATTCACTAGCTGATGAGA T..T C . . C T..T..C . . . T . . T . . c . r T. . C . . . . . . G . . .C . . . . . . . .C. . . . . T . r .T T . . . G r T . . C . r c . T T c c T A .... T .T. . . . . . G . C . . . . . . . .C. . T C T T T . . C . r . T.TC. . . . . T . . . .C. . . . . .T. . . C . . TC . . . . T . . C . . .G.C. . . . . . . .C. . . .T A . .C. . .G . . . .C. . AAACAAAATCCTCAGCCTAAATTAG TTTTGGTAACTTAACT-AAAAAGCGTCGAC G CTTGTAAGYCGAAGACTGGAG G G G G G G G G T C G G G T C GG - T C. . . GG T . CGT Cb6H 3 ' We sequenced a tiger shark from the Gulf of Mexico and found 46 nucleotide differences between our se- quence and that of Martin and Palumbi (1993). This difference is greater than that seen between all pairs of species in this study and is clearly too large of a difference to be explained by intraspecific polymor- phism. Because our observed restriction patterns for tiger sharks from the Atlantic and Pacific match pre- dictions made from our tiger shark sequence, our se- quence for the tiger shark is likely correct, and part of the sequence listed in Martin and Palumbi (1993) is not that of tiger shark. The remaining eight species — bignose, blacktip (C limbatiis), bull (C. leucas), dusky, silky (C falci- formis), scalloped hammerhead, great hammerhead, and lemon sharks — exhibited restriction patterns 58 Fishery Bulletin 97(1), 1999 identical to those predicted based on sequence data (Table 4). The list of species that showed no varia- tion in restriction pattern included silky and dusky sharks from widely disparate geographic locales. Discussion Restriction digests of the diagnostic 394-396 bp frag- ment proved a reliable way to distinguish among the Table 3 Predicted restriction fragments for seven enzymes based on nucleotide sequences. Species Alul Ddel Fokl Haelll HincW Hinfl Rsal Bignose 232, 84, 79 331,64 310, 85 205, 190 223, 148, 24 324, 71 251, 144 Blacktip 121, 111,79,43,41 333, 64 395 395 223, 148, 24 324, 71 251, 144 Bull 190, 121,43,41 331,64 395 395 223, 148, 24 324, 71 251, 144 Dusky 311,43,42 331,64 395 221, 174 371, 24 324, 71 395 Sandbar A 223, 79, 43, 41 331, 64 395 205, 190 223, 148, 24 324, 71 251, 144 Sandbar B 223, 79, 43, 41 331, 64 310, 85 205, 190 223, 148, 24 324, 71 251, 144 Sandbar C 223, 79, 43. 41 331. 64 310,85 205, 190 223. 148, 24 324, 71 251, 144 Silky 232, 79, 43, 41 331,64 310,85 221, 174 371. 24 324, 71 395 Spinner A 190, 121,43,41 331, 64 395 221. 174 371. 24 324, 71 395 Spinner B 190, 121,43,41 331,64 395 221. 174 371. 24 324, 71 251, 144 Lemon 190, 121.43,41 331.64 310,85 221, 174 371, 24 324, 71 395 "Tiger" A' 190, 121.43,41 331, 64 208, 187 221, 174 371. 24 216, 108, 71 251, 144 Tiger B 190, 121.43,28, 13 .331. 64 310,85 395 371. 24 395 395 Great Hammerhead 121, 111.85,89 331. 65 396 396 223. 149, 24 325, 71 396 Scalloped Hammerhead 311,83 256, 75, 63 394 221, 173 370, 24 323, 71 395 ' It was determined that this sequence was not that of a tiger shark See text Table 4 Number and location of sharks whose mtDNA was subjected to restriction enzyme digestion. Numbers in parentheses represent number of sharks from each source location in the sample. Species Bignose shark 4 Blacktip shark 7 Bull .shark 7 Dusky shark 7 Lemon shark 7 Sandbar shark 10 Silky shark 7 Spinner shark 7 Tiger shark 10 Great hammerhead 7 Scalloped hammerhead 7 Source of specimen Virginia. Atlantic Ocean (4) Virginia, Atlantic Ocean ( 1 >; Florida, Gulf of Mexico (6) Florida, Gulf of Mexico (7) Virginia, Atlantic Ocean (2); Florida, Atlantic Ocean (1 1; Australia, Pacific Ocean (3); Australia, Indian Ocean ( 1 ) Florida, Gulf of Mexico 1 6); Mexico, Gulf of Mexico ( 11 Florida, Gulf of Mexico (7); Hawaii, Pacific Ocean (3) Texas, Gulf of Mexico (3); Hawaii, Pacific Ocean (4) Virginia, Atlantic Ocean (1); Florida, Atlantic Ocean (3); Florida, Gulf of Mexico (1); Australia, Pacific Ocean (2) Florida. Gulf of Mexico (5); Australia, Pacific Ocean (2 1 Hawaii. Pacific Ocean (3) Florida, Gulf of Mexico, Florida (6); Florida, Atlantic Ocean (1) Virginia, Atlantic Ocean (1); Florida, Gulf of Mexico (2); Alabama, Gulf of Mexico (1); Texas, Gulf of Mexico ( 1 ); Campeche, Gulf of Mexico 1 2 1 Heist and Gold: Genetic identification of sfiarks 59 eleven species of carcharhiniform sharks. Each spe- cies possessed a unique set of restriction fragments, and even in the species where polymorphism was detected, misidentification was not a problem. This technique can be performed on small pieces of tissue collected at dockside and stored indefinitely at room temperature and moreover can distinguish between species whose carcasses may be difficult to discrimi- nate (Fig. 1). Similar DNA technology has been used for identifying carcasses in teleosts (Chow et al., 1993; Bartlett and Davidson, 1991 ), and for identifying plank- tonic fish eggs (Daniel and Graves, 1993) — cases where Figure 1 Ten-percent polyacrylamide gel of Alul digests in three species that produce very similar carcasses. Lane zero is a size standard, lanes one to three are sandbar sharks, lanes four to six are bignose sharks, and lanes seven to nine are dusky sharks. Numbers at left refer to sizes (base pairs) of size standards. Predicted sizes of fragments in each spe- cies are given in Table .3. morphological characters proved insufficient for spe- cies level identification . Use of genetic characters to identify species can be complicated by population structure. Although most marine fishes that are distributed across vast geographic stretches are open-ocean pelagics (Briggs, 1960), with presumed minor genetic differences across the range of a species (but see Crosetti et al., 1994; Graves et al., 1992), many large coastal sharks are distributed as multiple discrete populations (Compagno, 1984). It is thus important to know whether genetic differences among individuals are diagnostic of species or populations. Some questions, among others, are the follow- ing: to what degree are published genetic data useful in regions other than those from which the original specimens were collected, and can this method be used to identify populations as well as species? Baker et al. ( 1996), for example, were able to determine geographic origin of marine mammals they identified from flesh samples on the basis of genetic data. In our study, regional differences in restric- tion patterns were observed in only one species, spinner shark, whereas sequence differences that did not affect restriction patterns were observed in sandbar shark. In four species ( dusky, silky, scalloped ham- merhead, and tiger), sequences from one individual accurately predicted restriction sites in specimens collected thousands of kilometers from where the original speci- men was collected. The low level of intra- and interregional polymorphism within species is not sur- prising given the low genetic diversity typi- cally reported for sharks (Smith, 1986). In restriction fragment length polymorphism (RFLP) studies of whole mtDNA molecules. Heist et al. (1995, 1996) found a very low nucleotide sequence diversity of 0.036% in sandbar shark and 0.13% in Atlantic sharpnose shark (Rhizoprionodon terraen- ovae). The small numbers of substitutions between geographically distant popula- tions of sandbar and spinner shark ob- served in this study indicate that intraspe- cific diversity should not hinder species identification of these eleven species. The erroneous tiger shark sequence in the study of Martin and Palumbi (1993) indicates a further strength of the PCR- RFLP technique beyond use for forensic identification; it can be used to evaluate 60 Fishery Bulletin 97(1), 1999 validity of sequence data. Many molecular genetic or phylogenetic studies (or both) are based on se- quences sampled from only one individual of a spe- cies. Discovery of erroneous sequence data in such studies is becoming increasingly common (Derr et al., 1992; Helbig and Seibold 1996; Ledje and Arnason, 1996). PCR reactions are easily contami- nated by carryover DNA from other organisms or by other DNAs in the laboratory (Thomas, 1994). Mislabeling of tubes also can lead to incorrect as- signment of sequence data to species. We sequenced the entire cyt h gene in tiger shark from Florida and, in comparison with the "tiger shark" sequence re- ported by Martin and Palumbi (1993), found one nucleotide difference in the first (5'-most) 540 bp of the gene and 64 nucleotide differences in the remain- ing 606 bp of the gene. We hypothesize that the "ti- ger shark" sequence of Martin and Palumbi (1993) is a mixture that includes sequence data from an- other species. The discrepancy between published and newly determined sequences is similar to obser- vations on published bird sequence by Helbig and Seibold (1996) and on mammalian sequences re- ported by Ledje and Arnason (1996). In each case, the published sequences differed so greatly from newly obtained sequences that conclusions of the prior works were called into question. The free ex- change of sequence data through Genbank also fa- cilitates replication of errors. Validity of sequence data can, and perhaps should, be tested by amplify- ing DNA from several individuals of the same spe- cies, and then by determining whether predicted re- striction sites are present in all amplifications. Acknowledgments We thank Steve Branstetter, George Burgess, Mat- thew Callahan, Bob Hueter, Chris Jensen. Chris Lowe, Charlie Manire, and Craig Plizga for provid- ing tissue samples. Andrew Martin provided valu- able advice and assistance. This work research was supported by the MARFIN Program of the U.S. De- partment of Commerce (grant NA57FF0053). Part of this work was carried out in the Center for Bio- systematics and Biodiversity, a facility funded, in part, by the National Science Foundation (award DIR-8907006). Literature cited Baker, C. S., F. Cipriano, and S. R. Palumbi. 1996. Molecular genetic identification of whale and dolphin products from commercial market.s in Korea and .Japan. Mol. Ecol. 5:671-685. Bartlett, S. E., and W. S. Davidson. 1991. Identification of Thunnus tuna species by the poly- merase chain reaction and direct sequence analysis of their mitochondrial cytochrome 6 genes. Can. J. Fish. Aquat. Sci. 48:.309-317. Briggs, J. C. 1960. Fishes of worldwide (circumtropical) distribution. Copeia 1960:171-180. Broughton, R. E., and T. E. Dowling. 1994. Length variation in mitochondrial DNA of the min- now Cyprinella spiloptera. Genetics 138:179-190. Cabot, E. L., and A. T. Beckenback. 1 989. Simultaneous editing of multiple nucleic acid £md pro- tein sequences with ESEE. Comp. Appl. Biol. 5:233-234. Castro, J. I. 1993. A field guide to the sharks commonly caught in com- mercial fisheries of the .southeastern United States. U.S. Dep. Commer, NOAA Tech. Memo. NMFS-SEFSC-338. Chow, S., M. E. Clarke, and P. J. Walsh. 1993. PCR-RFLP analysis on thirteen western Atlantic snappers (subfamily Lutjanidae): a simple method for spe- cies and stock identification. Fish. Bull. 91:619-627. Compagno, L. J. V. 1984. FAO species catalogue, vol 4, part 1: sharks of the world. An annotated and illustrated catalogue of shark species known to date. FAO Fish. Synop. 125, 655 p. Crosetti, D., W. S. Nelson, and J. C. Avise. 1994. Pronounced genetic structure of mitochondrial DNA among populations of the circumglobally distributed grey mullet (Mugil cephalus). J Fish. Biol. 44:47-58. Derr, J. N., S. K. Davis, J. B. Wolley, and R. A. Wharton. 1992. Reassessment of the 16S rRNA nucleotide sequence from members of the parasitic Hymenoptera. Mol. Fhylogenet. Evol. 2:1.36-147. Daniel, L. B. Ill, and J. E. Graves. 1993. Morphometric and genetic identification of eggs of spring-spawning sciaenids in lower Chesapeake Bay. Fish. Bull. 92:254-261 Graves, J. E., J. R. McDowell, A. M. Beardsley, and D. R. Scoles. 1992. Stock structure of the bluefish Pomatomus saltatrix along the mid-Atlantic coast. Fish Bull. 90:703-710. Heist, E. J., J. E. Graves, and J. A. Musick. 1995. Population genetics of the sandbar shark iCarcharhinus plumbeus) in the Gulf of Mexico and Mid-Atlantic Bight. Copeia 1995:55,5-562. Heist, E. J., J. A. Musick, and J. E. Graves. 1996. Mitochondrial DNA diversity and divergence among sharpnose sharks. Rhizoprionodon terraenovae, from the Gulf of Mexico and Mid-Atlantic Bight. Fish. Bull. 94:664-668. Helbig, A. J., and I. Seibold. 1996. Are storks and new world vultures paraphyletic? Mol, Phylogenet. Evol. 6:31.5-319 Hoenig, J. M., and S. H. Gruber. 1990. Life-history patterns in the elasmobranchs: implica- tions for fisheries management. In H. L. Pratt Jr, S. H. Gruber, and T. Taniuchi (eds. I, Elasmobranchs as living resources: advances in the biology, ecology, systematics, and the status of the fisheries, p. 1-16. U.S. Dep. Commerce. NOAA Tech. Rep. NMFS 90. Holden. M. J. 1974. Problems in the rational exploitation of elasmobranch populations and some suggested solutions. In F. R. Harden-Jones (ed.). Sea fisheries research, p. 117- 137. John Wiley and Sons, New York, NY. Heist and Gold: Genetic identification of shiarks 61 1977. Elasmobranchs. In J. A. Gulland (ed.). Fish popu- lation dynamics, p. 187-214. John Wiley and Sons, New York, NY. IBI Mac Vector. 1991. IBI Mac Vector, 3.5 International Biotechnologies Inc. New haven, CT. Ledje, C, and U. Amason. 1996. Phylogenetic analyses of complete cytochrome b genes of the order Carnivora with particular emphasis on the Carniformia. J. Mol. Evol. 42:135-144. Martin, A. P. 1993. Application of mitochondrial DNA sequence analy- sis to the problem of species identification of sharks. U.S. Dep. Commer. NOAATech. Rep. NMFS 115:53-59. Martin, A. P., and S. R. Palumbi. 1993. Protein evolution in different cellular environments: cytochrome b in sharks and mammals. Mol. Biol. Evol. 10:873-891. Musick, J. A., S. Branstetter, and J. A. Colvocoresses. 1993. Trends in shark abundance from 1974 to 1991 for the Chesapeake Bight region of the U.S. Mid-Atlantic coast. U.S. Dep. Commer., NOAA Tech. Rep. NMFS 115: 1-18. NMFS (National Marine Fisheries Service). 1993. Fishery management plan for sharks of the Atlantic Ocean. U.S. Dep. Commer., Natl. Mar Fish. Serv. South- east Regional Office, St Petersburg, FL 33702, 167 p. Pratt, H. L., Jr., and J. G. Casey. 1990. Shark reproductive strategies as a limiting factor in directed fisheries, with a revision of Holden's method of estimating growth parameters. In H. L. Pratt Jr, S. H. Gruber, and T. Taniuchi (eds.), Elasmobranchs as living resources: advances in the biology, ecology, systematics, and the status of the fisheries, p. 97-109. U.S. Dep. Commer., NOAA Tech. Rep. NMFS 90. Sambrook, L., E. F. Fritsch, and T. Maniatis. 1989. Molecular cloning: a laboratory manual. Cold Spring Harbor Laboratory Press, Cold Spring Harbor, New York, NY. Smith, P. J. 1986. Low genetic variation in sharks (Chondrichthyes). Copeia 1986:202-207. Thomas, R. H. 1994. Analysis of DNA from natural history museum collections. In B. Schierwater, B. Streit, G. P. Wagner, R. Desalle (eds. ), Molecular ecology and evolution: approaches and applications, p. 311-321. Birkhauser Verlag, Basel. 62 Abstract.— Correlation analysis was used to investigate how effort and tem- perature changes influence lobster catches at different spatial and tempo- ral scales and how they may affect the use of catch statistics in stock assess- ments. At the largest scales examined (Atlantic coast of Nova Scotia, 50 yr), a significant correlation between catches and temperatures at short lags (0-3 yr) prior to 1974 suggests that catches were driven by temperature-induced changes in growth or lobster activity. However, changes in effort could not be ruled out as a cause of the observed cycles because sea surface temperatures may reflect weather conditions and the "fishability" of the grounds. Longer lags (6-8 yr) after 1974 are consistent with increased larval survival due to sharply rising temperatures and the "recruit- ment pulse" of the late 1980s. There was no clear relation between tempera- ture and catches at intermediate scales (statistical districts. 10 yr), but effort changes indicate that catches alone do not accurately reflect changes in lob- ster abundance. At the smallest scale examined (distances between ports, days) correlations of both temperature and effort with catch in areas with simi- lar coastal topographies indicated that the correlation between temperature and catch was not causative, i.e. changes in effort, were driven by wind events that were also influencing water tempera- tures. The results indicate that effort changes must be considered at all scales in stock assessments, but they become increasingly important at the smallest (i.e. <100 km. within years) scales. They also indicate that signifi- cant correlations between lobster catches and environmental parameters must be interpreted cautiously. Influence of temperature and effort on lobster catches at different temporal and spatial scales and the implications for stock assessments Peter Koeller Department of Fisheries and Oceans Invertebrate Fisheries Division Bedford Institute of Oceanography P.O. Box 1006, Dartmouth Nova Scotia, Canada B2Y 4A2 E-mail address koellerpigmardfo-mpogcca Manuscript accepted 19 March 1998. Fish. Bull. 97:62-70(19991. In the absence of suitable effort data, long-term, large-scale changes in American lobster (Homarus americanus) populations are gener- ally assessed by using summarized catch statistics (e.g. Harding et al, 1983; Pezzack, 1992; Drinkwater et al, 1991, 1996), under the assump- tion that catch trends reflect abun- dance changes. Two observations support this assumption. First, high exploitation rates in most areas probably remove the majority of animals recruiting to the fishery in any year (Miller et al, 1985). Sec- ond, similar catch trends over large areas of the Northwest Atlantic sug- gest a similar population response to a common environmental influ- ence (Pezzack, 1992). Given that this assumption is correct at these large scales, it is of fundamental interest in survey design and data aggregation decisions to determine the spatial and temporal scales at which this assumption holds. For example, can the catch statistic alone be used to assess changes in lobster populations between, say, adjacent ports within seasons? To understand the spatial and temporal variability of lobster catches on the Atlantic coast of Nova Scotia, Canada, and the un- derlying causes of this variability, 1 examined the influence of effort and an important environmental factor (temperature) on commercial lob- ster catches at different spatial and temporal scales. Temperature can affect lobster catches in the short term by altering the availability of lobsters to traps, i.e. by changing lobster behavior on the grounds (McCleese and Wilder, 1958) or by increasing growth of prerecruits to commercial sizes. In the long term it could also affect population growth by influencing survival of lar'v'ae or juvenile stages (Aiken and Waddy 1986). Changes in effort af- fect lobster catches directly but how these changes occur is complex. In a lucrative fishery with a relatively short (approx. 60 d) open season and restrictions on the number of traps, fishermen tend to take the maxi- mum number of lobsters possible with their gear. Consequently, changes in effort are more likely to be due to factors that decrease ef- fort. For example, fishermen could decrease the number of traps fished and frequency of hauls when catches are too low to make fishing worth- while. Because the fishery is con- ducted from small boats, adverse weather conditions will also de- crease effort. Conversely, fishermen may respond to increasing catch rates or good weather by fishing harder. Because catches alone are often used as an indication of stock status, it is important to under- stand how changes in effort may be influencing catches at the different spatial and temporal scales at which assessments are conducted. Koeller; Influence of temperature and effort cfianges on lobster catchies 63 \ ^ LFA31A 4i N"^ ♦ ~ 4 "-^ ^"^ 4, -c^ ^. -V "<^. <^\ * ^ "*"*. \ % ^\ \ ^^ <" ^ 'c. ^ ^%. >/^ \ '^ LFA31B LFA33 \ LEA 32 50 km Figure 1 Study area, including locations of ports along the eastern shore where fishermen provided catch and effort information. The number of index fishermen at each location providing data in 1994 is given in brackets after the port name. The dashed lines show the boundaries be- tween lobster fishing areas (LFAs) and statistical districts (SDsi. The location of ports that are discussed in the text are circled. Materials and methods Annual lobster catch is recorded by statistical dis- trict from purchase slips by the Department of Fish- eries and Oceans (DFO). These data are summarized by larger geographical units for assessment purposes, usually by lobster fishing areas (LFAs). In order to examine the large scale relationships between catches and temperature, catches were summarized in a similar manner for this analysis. The Atlantic coast of Nova Scotia (Fig. 1) was bisected into coast- lines of approximately equal length, named the east- ern shore (LFAs 31 A, 3 IB, and 32) and south shore (LFA 33). Long-term records of sea surface tempera- tures (SSTs) are available from Halifax harbour which is located on the boundary between these areas and can be considered representative of both. The three data series, i.e. total annual catches from the south and eastern shore and average annual SSTs, were smoothed with a 3-yr running average. Examination of the plots suggested that the series could be divided into two periods, a pre-1974 period of high-frequency, low-amplitude cycles, and a single post-1974 cycle of high amplitude. Correlations at lags of 0-10 years were then run for each period sepa- rately and combined. Daily catch and effort information is collected for DFO by selected index fishermen in some areas of Nova Scotia. At minimum these fishermen provide total lobster catches (usually weight in pounds) and the number of traps hauled on a daily basis during the fishing season. The number of index fishermen varies between areas but tends to cluster around key ports that are also sampled for length frequencies by port samplers. On the eastern shore, about 20 fish- ermen have kept log books along a 200-km stretch of coastline between Canso and Halifax, some begin- ning as early as 1986. In 1994 an additional 28 fish- ermen associated with the Fishermen's and Scien- tists' Research Society (FSRS) kept logs in this area. Most FSRS fishermen also deployed continuous tem- perature recorders (Vemco Ltd., Halifax, N.S.) on their traps during the 1994 fishing season, which extended from 20 April to 20 June in LFAs 3 IB and 32, and from 1 May to 31 June in LFA 31A. Because effort information is available only for a limited num- 64 Fishery Bulletin 97(1), 1999 Table 1 Statistics for correlation analysis of data series in Figure 1. Eastern shore South shore significant lag (yr) at lags (yr) maximum r r significant lags (yr) lag (yr) at maximun r r Pre- 197 4 Post-1974 1-3 2 — 8 0.683* 0.519 0-2 1 6 0.688* 0.532 * = significant at P<0.01 ber of index fishermen within each statistical dis- trict, total effort for each district was calculated by dividing its total catch by the average catch per trap haul (CPTH) for index fishermen in that district. Because seasonal trends were evident in tempera- ture, CPTH, and effort data, any analysis involving daily observations of these parameters were detrended by using first differencing i.e. /-(;'- 1), where / is the individual observation of temperature, CPTH, or catch. This avoided spurious negative cor- relations between the decreasing catches through- out the season that were due to depletion of the fish- able biomas, and increasing spring-summer tempera- tures. Because most fishermen on the eastern shore attempt to haul their traps daily and because longer soaks are usually associated with unfavorable weather (which tends to affect most fishermen in the same area ), there was no attempt to adjust for differences in soak times within or between each fisherman's time series. Inspection of the data showed that gaps in the indi- vidual time series tended to occur on the same days, which coincided with storm days and Sundays. Temperature information was also available from DFO moored recorders moored along the eastern shore during the period when logbooks were kept. Years with a strong association between tempera- ture and catch were identified for further study at smaller spatial and temporal scales as follows: daily, temperature, catch, and effort data from the 1986- 94 fishing seasons at the location with the most com- plete longer-term records, i.e. Port Bickerton, were detrended as described above and the resulting indi- vidual series correlated. This analysis identified 1991 and 1994 as years with significant correlations. The latter year was chosen for further investigation into the nature of this relationship because of the avail- ability of data at small spatial s.cales throughout the area from FSRS records. Three matrices of port ver- sus date, one each for the detrended temperature, CPTH and effort data, were then developed for fur- ther correlation analysis. Each matrix resulted in a port versus port matrix of correlation coefficients, i.e indices of similarity between port pairs in terms of the daily patterns of change for the respective pa- rameters. Correlations between port pairs could then be plotted against distance between them. Results and discussion Effort information is not available at the largest spa- tial and longest temporal scale examined (Atlantic coast of Nova Scotia, 50 years). At this scale, how- ever, there is a significant correlation between an- nual mean sea surface temperatures in the approxi- mate center of the area (i.e. Halifax harbor) and sum- marized lobster catches ( Fig. 2; Table 1 ) at short lags prior to 1974. This correlation was also noted by Campbell et al (1991). Catches are significantly cor- related (P<0.01) with temperatures for three cycles prior to 1974 at lags of 1-3 years on the eastern shore and 0-2 years on the south shore. After 1974, corre- lations are highest (although marginally insignifi- cant) at lags of 8 years on the eastern shore and 6 years on the south shore, approximately the time lobsters take to recruit to the fishery in these areas. This finding suggests that during this period, tem- perature or an unknown covariate also influenced the survival of larvae and subsequently increased recruitment to the fishery. The increase in catches in both areas after 1974 is consistent with the north- west Atlantic coast-wide "recruitment pulse" ( Pezzack 1992 ), which is generally attributed to a com- mon environmental influence, although Drinkwater et al ( 1996) concluded that this environmental influence was not temperature. Prior to 1974 the near in-phase changes in temperature and catch along the eastern and south shore could be due to temperature-induced changes in catchability during warmer years i.e. to greater activity of lobsters on the grounds or faster growth to legal size, or to both. However, without long- term effort information, changes in effort cannot be Koeller: Influence of temperature and effort changes on lobster catches 65 ruled out as a cause of the pre-1974 catch cycles. For example, because Halifax harbour sea surface temperatures reflect weather con- ditions in the study area, years with colder sea surface temperatures could also be years with less fishing activity. It is noteworthy that along the eastern shore both pre- and post- 1974 catches lag temperatures by 1-2 yr longer than on the south shore. Hudon ( 1994 ) noted that the population along the eastern shore may be growth-limited owing to colder water temperatures than those in adjacent areas, and one would expect this area to re- act more slowly to the more moderate tem- perature increases. Catch and effort in smaller statistical dis- tricts along the eastern shore, along with available temperature indices in the area (Fig. 3), reflect the catch patterns of most other lobster fishermen in the northwest Atlantic during this period, i.e. peak catches in the late 1980s and subsequent declines. There was an increase in effort along the eastern shore throughout the period. There is no apparent relation between catches and available temperature indices, although the warmest water and highest catch both oc- curred in the same year ( 1989). Port Bickerton reflected the catch and effort pat- tern of the eastern shore as a whole, including an increase in effort throughout most of the period ( Fig. 4). Note that the increase in the average daily num- ber of trap hauls leveled off when catches started to decline after 1991. Perhaps fishermen were increas- ing effort in response to increasing lobster catches during the 80s. Pringle and Duggan ( 1984) noted that as many as 25% of the maximum number of traps (eastern shore trap limit is 250) can remain unused in the Nova Scotia lobster fishery. Although the use of catches alone would have overestimated the rate of increase in the population during the 80s and under- estimated the subsequent decline, the differences are small and would not have affected conclusions on an- nual population changes. A significant correlation between average daily temperatures and CPTH at Port Bickerton occurred in only 2 of the 8 years for which temperature, catch, and effort information is available at this scale, sug- gesting that short-term, temperature-induced changes in catchability occurred, or were detectable, during those years (Tables 2 and 3). However, a mul- tiple regression of daily catch (dependent variable) versus effort and temperature during these years also indicates that short-term changes in lobster catches were largely determined by effort changes. Cross 9.5 9 8.5 8 7.5 7 6.5 6 Eastern shore temperature , t700 Figure 2 Total annual lobster catch for the eastern shore and south shore of Nova Scotia and average annual sea surface temperatures in Halifax harbor, 1950-92. All data series were smoothed with a three year average. Table 2 Pearson correlation coefficients for average daily tempera- ture (at 10 m) vs. total daily catch, total daily effort, and average daily catch per trap haul for index fishermen at Port Bickerton 1986-94. Catch Effort (trap hauls) Average CPTH 1986 0.0491 0.1066 0.0830 1987 0.2158 0.0065 0.1740 1988 -0.0380 -0.0373 0.1060 1989 -0.0551 0.0718 0.1580 1990 0.1317 0.1239 0.1690 1991 0.311* 0.1014 0.408* 1992 -0.1426 -0.2853 0.2110 1993' 1994 0.2477 0.1888 0.262* ' = significant at P<0.05 ' Temperature data for 1993 were not available in Bickerton dur- ing the fishing season. correlations between total daily catch and effort at Port Bickerton at lags of 0-7 days showed no signifi- cant correlations at lags other than 0, indicating that fishermen's effort was not influenced by catches at short time scales. 66 Fishery Bulletin 97(1), 1999 Halifax SST (Jan-Mar) Halifax SST(Apr-Jun) Bickerton (1 Im) (N m 0^ ON O On Figure 3 Total annual lobster catch (topi and total annual effort (middlel for east- ern shore statistical districts from 1986-95. the years for which effort information is available from index fishermen. Also shown for these years are three temperature indices (bottom), including average sea surface temperatures (SST I in Halifax harbour from January to March; average SST from April to June i.e. during the lobster season; and average water temperatures (11 m) at Port Bickerton during the lobster season. The bars show the average temperature for the first (lower) and last (upper) week of the lobster season. Correlation coefficients of the daily first differenced CPTH data from individual fishermen on the eastern shore were plot- ted against distance from Port Bickerton and Three Fathom Harbour (Fig. 5), situ- ated about 125 km apart near the north- eastern and southwestern end of the study area (Fig. 1). Correlations were strongest in the immediate vicinity of these ports, decreased to insignificance at about 50 km, then increased to a second maximum at 125 km. This pattern was still apparent when all ports were in- cluded in the analysis (Fig. 5, bottom), but the least-squares fit to the polynomial was not as good, perhaps because local factors influence the main spatial trend in other areas. In contrast, short-term changes in water temperature are strongly correlated throughout the area (Fig. 6) because they are caused by wind events that impact the eastern shore as a whole. Apparently, the correlation of catches and temperatures at Port Bickerton in 1994 (Table 2) oc- curred only at either end of the eastern shore. To determine if this correlation was truly a direct environmental influence on catch, I plotted the daily correlation coef- ficients of all ports against distance sepa- rately for CPTH and effort (Fig. 7). The fourth-order polynomials fitted to the data are similar for both, indicating that the similarity in catches between ports on ei- ther end of the eastern shore and their similarity to short-term temperature changes are mainly due to similarities in the daily pattern of change in effort. This is confirmed by plotting the coefficients for effort versus temperature directly. Similar results were obtained for 1991, the other year in which significant corre- lations between temperature and catches were found at Port Bickerton. The reason why fishermen separated by 125 km have similar day-to-day fishing patterns is ap- parent when one examines the nature of the coastline along the eastern shore (Fig. 1 1. The coast near Three-Fathom Harbour and Bickerton is characterized by rela- tively long and narrow bays generally ori- ented in a north-south direction, which offer more shelter than the relatively ex- posed coastline between them. If the winds that cause temperature decreases are also those that make fishing more difficult, Koeller: Influence of temperature and effort changes on lobster catches 67 210 - r ■/< ./v A 200 - 180 > S. 200 - 3 '/ ^ xA-" ^ ^ 160 - 140 Aver erage S- 190 * . , '/ -/ N O.0Q t 180^ ^ - 120 B. f ^ y^^ \ ^ 100 daily catch ( ly CPTH (lb •a S" 170 / TRAPS CATCH ■ ■ - 80 - 60 < 160 { - - - CPTH - 40 - 20 0 X N^ OU H 86 87 88 89 90 91 92 93 94 95 96 Year Figure 4 Annual averages of daily catch, trap hauls, and catch per trap haul for ndex fishermen at | Port Bickerton 1986-96. Table 3 Multiple regression analysis of total daily catch (dependant variable), total daily effort (trap hauls) and average daily tempera- tures for the two years where significant correlations between temperature and catch per trap haul were found (Table 1 ). Statis- tics shown include slope and intercept values (B), standard error of B, and regression coefficients (Beta), with their significance levels (Sig. T). B SEB Beta T Sig.T 1991 Effort Temperature (Constant) 1.1382 50.9725 -0.9795 0.1055 62.0185 34,8649 0.8241 0,0628 -0,0280 10.7840 0.8220 0.9777 0.0000 0.4148 1994 Effort Temperature (Constant) 0.4000 61.3956 -10.4788 0.0528 51.8943 28.9416 0.7179 0.1121 -0.3620 7.5760 1.1830 0.7188 0.0000 0.2423 e.g. those that are southerly and enter directly into the bays, and are of a minimum force, or both, one would expect a positive correlation between water tempera- tures and catches in these areas. The apparently spu- rious correlation between lobster catches and tempera- ture would be weaker in the exposed central area where fishing activity is vulnerable to winds from a wider di- rection and of less force than those which significantly impact near-shore water temperatures. Although it is clear that the observed correlations between short-term changes in lobster catches and temperatures are strongly influenced by changes in effort, these results do not rule out a concurrent direct affect of tempera- ture on lobster activity and catches, i.e. effect of catchability. Possibly fishermen are aware that winds of a certain direction and force affect lobster activity and catchability and alter their fishing activity accord- ingly, or the factors that decrease fishing activity also decrease lobster catchability independently. These results indicate that although changes in effort can influence lobster catches at all spatial and temporal scales examined, data on effort is particu- larly critical in correctly interpretating the meaning of changes in catch at smaller scales i.e. <100 km and <1 yr. In particular, they indicate that small- scale correlations between environmental param- eters such as winds and water temperatures with lob- ster catches alone should be treated cautiously because inferences made about their underlying mechanisms may be wTong. For example, Hudon (1994) concluded that differences in annual catches between approxi- mately 100-km lengths of the Nova Scotia coastline were due to differences in the orientation of these coast- lines to upwelling winds, the resultant change in wa- 68 Fishery Bulletin 97(1), 1999 -0.2 -0.4 R^ = 0.6506 50 100 150 200 Distance from Three Fathom Harbour (km) 1 ■^ ^ o.x c u 0.6 0.4 o u c 0.2 o ^ 0 ty> it OS) -0.2 v iC -0.4 00 o 0 50 100 150 Distance from Bickerton (km) R- = 0.117 50 100 150 Distance from port (km) 200 Figure 5 Pearson correlation coefficients of first differenced daily CPTH of all index fishermen along the eastern shore versus increasing dis- tance from Three Fathom Harbour (top) and Bickerton (middle). The bottom figure plots the same statistic for all ports versus dis- tance from each other. A fourth order polynomial provided the best least-squares fit to each data set. ter temperature, and the effect of the latter on lobster productivity. This study suggests that at least some of these differences may be due to differences in effort and "fishability" of the coastlines involved. These results also give a general indication of the sampling effort (i.e. number of index fishermen) and distribution (spacing of index fishermen) required of a program whose objective is to characterize and as- sess changes in effort for an entire coastline. Because changes in effort from index fishermen separated by short distances, i.e. within the working reaches of a fishing port, are strongly correlated, it would be more effective for any given total sampling intensity to space samples evenly along that coastline, rather than cluster them in ports. Since the correlation of index fishermen's catch data decreases quickly at dis- Koeller: Influence of temperature and effort changes on lobster catches 69 ■south u o O 4 - 0 . I I . II I I I M i I I ' I I I I I ' M i I II I ; ! I n I I I I II II I I I I I I I H ! I I I M I l-H o. < o. < 4 o. < 2 S A 03 n A 2 2 2 S O -4 00 (N 6 1 0.9 -»^ 0.8 c: u 0.7 o 0.6 u y 0.5 c •s 0.4 O.H ^2 (3 0.2 0.1 0 — I 1 1 — 50 100 150 Distance from recorder (km) 200 Figure 6 (Top) Average daily temperatures from temperature recorders deployed by in- dex fishermen on their traps during 1994. for the southern (south of Murphy Cove), middle (Murphy Cove to Liscomb) and northern (north of Liscomb) sec- tions of the eastern shore. (Bottom) Pearson correlation coefficients of first differenced temperature data from all temperature recorders deployed by in- de,x fishermen versus distance between instruments in 1994. tances greater than about 25 km, the samples should be more closely spaced. On the eastern shore, where the annual average number of fishermen providing information is about 20, the resulting 10-km spac- ing at this sampling level should be adequate to pro- vide a representative sample. Although these results have not eliminated the possibility of temperature-induced changes in catchability, they do indicate that quantifying these changes with fishermen's data would be extremely difficult. Acknowledgments I wish to thank the Nova Scotia lobster fishermen who provided information used in this study, espe- cially those who have collected data since the begin- 70 Fishery Bulletin 97(1), 1999 1.0- t o U 1.0 n 0.5- 0.1 - 1994 Effort versus temperatiire 200 Distance from port (km) 100 200 Figure 7 Fourth order polynomials as shown in the bottom of Figiire 5, but for catch per trap haul and trap hauls separately, for the years in which significant correlations of daily tempera- ture and catch per trap haul were found at Port Bickerton (199 1 , 1994 ). Also shown is the polynomial curve for individual fishermen's daily effort versus daily average tempera- tures in their area in 1994. Individual points have been omitted for clarity, but their dis- tribution is similar to the bottom of Figure 5. ning of DFO's lobster logbook program. Their con- tinuing efforts increase the value of this database each year. The cooperation of Randy Baker, Sean Smith, and fishermen of the Fishermen's and Scien- tists' Research Society is gratefully acknowledged for access to additional data for the 1994 season, includ- ing temperatures. Ed Verge provided historical tem- perature data collected by the Physical and Chemical Sciences Branch, DFO, and Ron Duggan summarized much of the data for analysis. Bob Roger and Steve Smith provided statistical advice, and John Tremblay and Doug Pezzack reviewed early versions. I am par- ticularly indebted to three anonymous reviewers whose suggestions significantly improved the manuscript. Literature cited Aiken, D. E., and S. L. Waddy. 1986. Environmental influence on recruitment of the american lobster 0.15) to determine which independent variables accounted for most of the variation in pre- dictive equations (Zar, 1984; Neter et al., 1985). The resulting regression models were then used to pre- dict WC and calculate percentage error (PE): PE = {(|actual - predicted] / actual) x lOO}. Finally, we calculated percentage dry weight from both actual (determined from drying) and predicted (from regression equations) water content and com- pared the variation between these values to the range of percentage dry weights commonly observed for yellow perch and alewife (Tables 1 and 2). Results Significant (P<0.05 ) positive relationships were found in simple linear regressions between the dependent variable WC and the independent variables TOBEC, WWT, and TL for both yellow perch and alewife (r-^=0.66 to 0.99). Simple linear regressions oi TOBEC on WC produced r'^ values of 0.933 and 0.667 for yel- low perch and alewife, respectively. Multiple linear regressions with two independent variables (WWT and TOBEC) gave excellent fits to the data for both species. The equation for yellow perch was WC = 3.46239 + 0.69844 x WWT + 0.00559 x TOBEC (r-=0.998). Lantry et al.: Electrical conductivity to estimate water content of Perca flavescens and Alosa pseudoharengus 73 Table 1 Yellow perch individual water content (gl from measurements laet) and predicted (pred) from regression functions. TOBEC is the average scanner conductivity index (based on five consecutive measurements). "^DWTis the percentage dry weight and is equiva- lent to: |( I wet weight - water I )/wet weight) x 100. Values in parentheses are percentage errors (PEs) referring to water (g) and '/f'DWT, and are equivalent to l( I actual - predicted I )/actual| x 100. Total length (mm) Wet weight (g) TOBEC Water •. 100. Values in parentheses are percentage errors (PEs) referring to water (gl and %DWT and are equivalent to |( | actual - predicted | )/actuall x 100. Predicted (pred) values are from the regression function with wet weight (g) and TOBEC as the independent variables. Total length (mm) Wet weight (g> TOBEC Water (g) (act) Water 0.80) and fish (/•2=0.67 to 0.988). In fish however, TOBEC predicted total wet weight equally well. If more than just test animal size affects conductivity readings, then evidence of changes in the total body content of electrolytic salts should also be apparent when TOBEC and water content values are divided by wet weight. Wet-weight-standardized TOBEC and wa- ter content values were not related for alewife, sun- shine bass, or red drum (Fig. 2). The apparent trend in the yellow perch data is counterintuitive (Fig. 2) to the expected trend of increased conductance with increased water content. The strength of predictive equations for fish in the above four data sets may be solely due to effects from fish size (e.g. serum and cellular fluid volumes). The expected relationship for these parameters is apparent in our plot of the test data set from Jaramillo et al. ( 1994; Fig. 2). The nu- trient content of diets fed to fish in those experiments was carefully controlled within groups and varied only between groups. Evidence from our study indi- cates some promise for using TOBEC for fish in situ- Lantry et al.: Electrical conductivity to estimate water content of Perca flavescens and Alosa pseudoharengus 77 Sunshine bass ♦ n. 71 (Brown et al 1^3) 70- ♦ #»♦♦ % 69 ♦ ♦a ♦ 6K ♦♦ ♦ ♦ ♦ ♦ ♦ 67 ♦% ♦ ♦ I 66 H . 65- 1 — -* K- -^U ♦ 1 82 80f fBaietal. 1994) 78 76 74 72 70 68 Red dnim * * • ♦ ♦ *« t 0.75 0.95 1.15 1.35 1.55 0.75 1.15 1.55 1.95 2.35 81 76 71 66 61 56 Chaimel catfish . (JaramiUo et al. 1994) ♦ ♦♦ ♦ ♦ ••♦.. 1 1- 1 1 0.5 1.0 1.5 TOBEC/g 2.0 Figure 2 Comparisons between standardized values of percentage water (water content/wet weight) and total-body electrical conductivity (TOBEC/wet weight) for yellow perch, alewife, sunshine bass, red drum, and channel catfish. Data for sunshine bass, red drum, and channel catfish are from Brown et al. (1993). Bai et al. (1994). and Jaramillo et al. (1994). ations where nutritional status can be controlled (e.g. in aquaculture) and electrolytic balance is given suf- ficient time to equilibrate throughout all bodily fluid (i.e. serum, cellular, and extracellular) compartments. Our ultimate goal was to use our predictions of water content to assess the energy content of yellow perch and alewife. Our energy density relationships (Rand et al., 1994; Lantry, 1997) use percentage dry weight as the independent variable. Percentage dry weight values calculated from predicted water con- tent did not, however, correspond to values calculated from measured water content for any of the five fish species used in TOBEC studies (Fig. 3). Our analy- sis indicates that further evaluations of the use of TOBEC to predict fish body composition are war- ranted. Fish size should be constrained to narrow ranges, and percentage water ((water content / wet weight) X 100) between individuals of different con- dition should be evaluated. By controlling fish size, conductivity differences due to body geometry could 78 Fishery Bulletin 97(1), 1999 D 35 30 25 1- 20 15 10 Yellow perch Perfect correspondence 36 34 32 30 28 15 20 25 Sunshine bass 30 35 26 Perfect correspondence 26 28 30 32 34 36 38 Channel catfish 36 34 32 30 28 26 24 2^ 35 30 25 20 15 10 14 16 18 20 22 24 26 Red drum Perfect correspondence ^ 20 25 30 35 2 24 26 28 30 32 34 36 38 Actual %DWT Wet weight and TOBEC Wet weight Figure 3 Comparisons between actual and predicted values of percentage dry weight (dry weight/wet weight) for yellow perch, alewife, sunshine bass, red drum, and channel catfish. Data for sunshine bass, red drum, and channel catfish are from Brown et al. ( 1993), Bai et al. ( 1994), and Jaramillo et al. ( 1994). The line in each panel represents the location of equivalence between actual and predicted. be reduced and wet weight could be eliminated from prediction equations. Variability associated with the scanning equip- ment, fish preparation prior to scanning, fish condi- tion, and geometry offish within the scanner cham- ber may generate errors too high to accurately pre- dict ecologically significant changes in fish body com- position. Analysis of TOBEC readings taken in fish fed, fasted, frozen, and thawed in Bai et al.'s study (1994) indicates that both physiological and physi- cal states affect conductivity values. Dehydration in terrestial animals has also been obsen'ed to cause disproportionate changes in TOBEC values (Wals- berg, 1988). Our fish were frozen in water and prob- ably were not dehydrated; however, the death of the fish and the effect of freezing and thawing may have altered conductance. Also, the potential existed for exchange between body water and the water sur- Lantry et al.; Electrical conductivity to estimate water content of Perca flavescens and Alosa pseudoharengus 79 rounding the fish during fi-eezing and thawing. The measurement of TOBEC values, however, did not pro- duce consistent improvement in the predictabihty of whole-body water content in fish from any of the five data sets considered. This analysis indicates that TOBEC procedures will not be able to predict with suf- ficient accuracy the water content offish sampled fi-om field situations and fi-ozen in water for later analysis. Acknowledgments Funding for this research was provided for under grant number SFI/EPRI 92-03 from Electrical Power Research Institute and Niagara Mohawk Power Cor- poration. We extend our gratitude to Guey Wong Shu and Christine Morris for assisting in collection and processing alewife and we thank John Forney, An- thony VanDeValk, and Thomas Brooking for assis- tance in collection of yellow perch. We acknowledge Cornell University and the staff at the Cornell Uni- versity Biological Field Station at Oneida Lake for providing equipment for sampling and facilities for processing yellow perch samples. Finally, we thank Lars Rudstam, John Forney, Mark Olson, and Chris- tine Mayer for their reviews of the original manuscript. Literature cited Bai, S. C, G. R. Nematipour, R. P. Perera, F. Jaramillo Jr., B. R. Murphy, and D. M. Gatlin III. 1994. Total body electrical conductivity for nondestructive measurement of body composition of red drum. Prog. Fish-Cult. 56:232-236. Bracco, E. F., M.-U. Yang, K. Segal, S. A. Hashim, and T. B. Van Itallie. 1983. A new method for estimation of body composition in the live rat. Proc. Soc. Exp. Biol. Med. 174:143-146. Brown, M. L., D. L. Gatlin III, and B. R. Murphy. 1993. Non-destructive measurement of sunshine bass, Morone c/j rvsops (Rafinesque) x Morone saj:a?;7is (Walbaum), body composition using electrical conductivity. Aquae. Fish. Manage. 24:585-592. Castro, G., B. A. Wunder, and F. L. Knopf. 1990. Total body electrical conductivity (TOBEC ) to estimate total body fat of free-hving birds. Condor 92:496-499. Craig, J. F. 1977. The body composition of adult perch, Perca fluviatilis in Windermere, with reference to seasonal changes and reproduction. J. Anim. Ecol. 46:617-632. Domermuth, W., T. L. Veum, M. S. Alexander, H. S. Hedrick, J. Clark, and D. Eklund. 1976. Predictions of lean body composition of hve market swine by indirect methods. J. Anim. Sci. 43:966-976. EM-SCAN Inc. 1991. EM-SCAN Model SA2 Small Research Animal Body Composition Analyzer operation manual. EM-SCAB Inc., Springfield, IL, 64 p. Fiorotto, M. L., W. J. Cochran, R. C. Funk, H. Sheng, and W. J. Klish. 1987. Total body electrical conductivity measurements: ef- fects of body composition and geometry. Am. J. Physiol. 252:R795-R800. Hartman, K. J., and S. B. Brandt. 1995. Estimating energy density of fish. Trans. Am. Fish. Soc. 124:347-355. Jaramillo, F., Jr., S. C. Bai, B. R. Murphy, and D. M. Gatlin III. 1994. Application of electrical conductivity for non-destruc- tive measurement of channel catfish, Ictalurus punctatus, body composition. Aquat. Living Resour. 7:87-91. Keim, N. L., P. L. Mayclin, S. J. Taylor, and D. L. Brown. 1988. Total body electrical conductivity method for estimat- ing composition: validation by direct carcass analysis of pigs. Am. J. Clin. Nutr. 47:180-185. Lantry, B. F. 1997. Bioenergetic allometries of percids and gizzard shad: implications for estimating predation on the changing prey assemblage in Oneida Lake, NY. Ph.D. diss.. State Univ. New York College of Environmental Science and Forestry, Syracuse, NY, 123 p. Neter, J., W. Wasserman, and M. H. Kutner. 1985. Applied Linear Statistical Models, 2nd ed. Irwin, Homewood, IL, 1127 p.. Presta, E., K. R. Segal, B. Gutin, G. G. Harrison, and T. B. Van Itallie. 1983. Comparison in man of total body electrical conduc- tivity and lean body mass derived from body density: vali- dation of a new body composition method. Metab. Clin. Exp. 32:524-527. Rand, P. S., B. F. Lantry, R. O'Gorman, R. W. Owens, and D. J. Stewart. 1994. Energy density and size of pelagic prey fishes in Lake Ontario, 1978-1990: implications for salmonine energetics. Trans. Am. Fish. Soc. 123:519-534. Rand, P. S., D. J. Stewart, B. F. Lantry, L. G. Rudstam, O. E. Johannsson, A. P. Goyke, S. B. Brandt, R. O'Gorman, and G. W. Eck. 1995, Effect of lake-wide planktivory by the pelagic prey fish community in Lakes Michigan and Ontario. Can. J. Fish. Aquat. Sci. 52:1546-1563. Stewart, D. J., and F. P. Binkowski. 1986. Dynamics of consumption and food conversion by Lake Michigan alewives: an energetics modeling synthe- sis. Trans. Am. Fish. Soc. 115:643-661. Stewart, D.J., D. Weininger, D. V. Rottiers and T. A. Edsall. 1983. An energetics model for lake trout, Salvelinus namaycush: application to the Lake Michigan population. Can. J. Fish. Aquat. Sci. 40:681-698. SYSTAT Inc. 1993. SYSTAT 5.03 for Windows. Evanston, IL. 750 p. Walsberg, G. E. 1988. Evaluation of a nondestructive method for determin- ing fat stores in small birds and mammals. Phys. Zool. 61:153-159. Zar, J. H. 1984. Biostatistical analysis, 2nd ed. Prentice Hall, Engle- wood Cliffs, NJ. 80 Abstract.— We measured age and growth of larval striped bass {Morone saxatilis) and white perch (M. ameri- cana) and tested whether growth and survival were enhanced in relation to a seasonal pulse ("bloom") of high zoo- plankton abundance. Growth rates were lowest before the zooplankton bloom and highest afterwards for both fish species. An index of recruitment po- tential (instantaneous growth rate. G. divided by instantaneous mortality rate, Z) did not relate clearly to either water temperature or to zooplankton abundance in the case of striped bass but did relate to both factors for white perch. Retrospective analysis of hatch dates in recruited juvenile striped bass from the same year class indicated that later, faster growing cohorts were un- der-represented when compared to the larval cohort distribution, and that co- horts that co-occurred with high densi- ties of the cladoceran zooplankton Bosmina freyi were over-represented. Comparison of these results with simi- lar analyses from other systems sug- gests that biotic controls on year-class strength may predominate in estuarine systems where physical factors are rela- tively damped (Hudson) but may play relatively minor roles in those systems with high physical variability. Growth, mortality, and recruitment of larval Morone spp. in relation to food availability and temperature in the Hudson River Karin E. Limburg Michael L. Pace Institute of Ecosystem Studies, Millbrook, New York 12545 Present address (for K Limburg) Department of Systems Ecology University of Stockholm S-106 91 Stockholm, Sweden E-mail address (for K Limburg) Karin_L(g'system ecology su se Kristin K. Arend Oberlin College Oberlin, Ohio 44074 Manuscript accepted 1 April 1998. Fish. Bull. 97:80-91(1999). Survival (during the first year of life determines the year-class strength of many fish. The degree to which biotic or physical factors regulate recruitment of fish from larval to juvenile stages varies from one aquatic system to another, and even temporally within systems (Leggett and DeBlois, 1994). The factors pro- moting or inhibiting recruitment are a subject of intense interest both in theoretical analysis and practical management offish populations (cf Hilbom and Walters, 1992), Mortal- ity of young fish below harvestable size is difficult to observe and diffi- cult to measure by means of standard approaches to population assessment (e.g. mark and recapture). The age and growth rates of young fish, however, can be pre- cisely quantified with otolith analy- sis (cf. Campana and Neilson, 1985; Secor et al., 1995). This technique has permitted the determination of hatching dates, age, and growth rates of both larvae and juveniles of many species. Comparisons of hatching dates and gi'owth between life stages (e.g. larvae vs. juveniles) allow inferences about the fates of larvae. For example, if larval sur- vival is particularly high during a specific time period, analysis of ju- venile hatch dates should reflect the differential survival (assuming ad- equate sampling). These measures can also be used with repeated sam- plings of the population to track the growth and mortality of specific age classes (e.g. weekly cohorts) over time. Relationships between these cohort-based rates can be analyzed with respect to variables such as food availability, predation, and the physicochemical environment. We previously documented that larval striped bass (Morone saxa- tilis) and white perch (M. ameri- cana), which co-occurred with in- creases in crustacean zooplankton in the tidal Hudson River, showed a potential energetic advantage compared with larvae that preceded the zooplankton increase (Limburg et al., 1997). A major increase ("bloom") in zooplankton in early summer is a key feature of the Hudson estuary. Larvae occurring before the bloom are exposed to sparse zooplankton densities and low temperature, both of which are associated with mortality risks I Rogers and Westin, 1981; Chesney, 1989; Margulies, 1989; Uphofif, 1989; Tsai, 1991; Cowan et al., 1993 ). Larvae occurring during and after the bloom have high consump- LImburg et aL: Growth, mortality, and recruitment of larval Morone spp. 81 tion rates and are at lower risk of exposure to subop- timal or lethal temperatures. Do these temperature and food conditions trans- late into differential survival conditions for larval Morone spp. in the Hudson River? We addressed this question using methods of otolith age and growth- rate reconstruction. Specifically, we tested if periods of high food availability (zooplankton bloom condi- tions) were also periods of high larval growth rates. Further, by analyzing the otoliths of juveniles of the same year class (i.e. successful recruits from the lar- val stage), we also assessed whether or not fish that survived the larval period had high larval growth rates and hatching dates associated with the zoop- lankton bloom. Studies of Morone saxatilis in Chesapeake Bay tributaries have used both interannual variations in larval abundance (Uphoff, 1989) and cohort analy- sis methods (Rutherford and Houde, 1995; Secor and Houde, 1995; Rutherford et al., 1997) to estabUsh that rapid drops in water temperature below 12°C virtually eliminate larval cohorts. These studies con- cluded that the frequency of low temperature events can be an important factor in the recruitment of striped bass in those systems. In contrast. Pace et al. (1993) found no evidence that temperature af- fected interannual variation of Morone spp. recruit- ment in the Hudson River, although Dey ( 1981) pre- sented evidence for temperature-induced mortality in Hudson River striped bass in 1976 (temperature dropped suddenly from 15° to 12°C in late May). Thus temperature effects, although not detectable among years, might still manifest themselves as seasonal variability. In this paper we examine the 1994 year classes of striped bass and white perch, including an appraisal of the relation of zooplankton abundance and temperature to growth and mortality of indi- vidual cohorts. With this information, we can make tentative comparisons between analyses for the Hudson River and Chesapeake Bay. Materials and methods The Hudson River, a partially stratified estuarine- river system in New York State, is tidal up to the Green Island dam, 245 river kilometers (rkm) from the mouth. All life stages of white perch are found throughout the oligohaline and freshwater sections of the estuary (Klauda et al., 1988). Striped bass adults migrate into the estuary in spring to spawn. Larvae are found in both freshwater and oligohaline sections (Boreman and Klauda, 1988). Juveniles spread throughout the estuary and move seaward during late summer and fall (Dovel, 1992). Larval collections and preparation Field collections of larval Morone spp. were under- taken in the spring and early summer of 1994. Sam- pling details are described in Limburg et al. (1997). Briefly, three sites, in the upper (Kingston, rkm 148), middle (New Hamburg, rkm 105), and lower (Haverstraw Bay, rkm 65-70) portions of the estu- ary were sampled between 18 May and 6 July; all sites overlapped with spawning areas. Sampling fre- quency was designed to examine larval and zooplank- ton d3Tiamics before, during, and after the intensive bloom of the cladoceran zooplankton Bosmina freyi i=Bosmina longirostris with older nomenclature). Fish were collected during the daytime with 5-min oblique tows of 0.5-m diameter bongo nets (500-mm mesh). Five replicate tows were conducted at each site on each date. Samples were preserved in 100% ethanol which, when diluted with the sample, reached a final concentration of no less than 75%. Care was taken not to over-dilute the samples below this concentration because of risk of otolith erosion (Brothers, 1987). In the laboratory, fish were sorted by family and stored in fresh 100% ethanol. Three replicates were chosen at random from the five available at indi- vidual stations for each sampling date, and up to twenty Morone individuals were arbitrarily selected from each replicate; more replicates were used when larval abundances were low. The larvae were re- hydrated, soaked for 15 min in sodium borate buffer solution (30% saturated sodium borate), and then cleared by adding a small quantity of trypsin to the buffer solution. Morone larvae were measured to 0.1 mm notochord length (NL) and identified to species by a combination of internal and external charac- ters (Limburg et al., 1997). Lengths were not cor- rected for preservation shrinkage, but rehydration and clearing fully relaxed constricted musculature so that lengths were readily measured. Otoliths were visible under 25x magnification on a stereomicro- scope. Fine insect pins, the tips of which were sharp- ened further, were used to dissect the sagittal otoliths. Otoliths of fish less than 5 mm NL were cleaned in deionized water and placed directly in mineral oil for later viewing. Following the sugges- tion of Rutherford (1992), otoliths of larger (>5 mm NL) fish were embedded in Spurr's epoxy, then ground and polished with a series of increasingly fine grinding papers (down to 3 mm). Juvenile fish collection and preparation In order to make comparisons between recruited ju- veniles and larvae, we obtained a total of 218 juve- 82 Fishery Bulletin 97(1), 1999 nile striped bass from five sites along the Hudson River (at river kilometers (rkm) 40, 63, 95, 114, and 201 ) during July (n =37), August («=88), and Septem- ber (?! =93) ft-om state-run fishery surveys. Fish were collected either with a 30.5-m (lOO-ft) beach seine with 6.1-m bag, 0.64-cm mesh, or with a 61-m (200- ft) beach seine with 1.27-cm stretch mesh (0.64 cm square). Fish were held on ice in the field, then stored ft^ozen. In the laboratory, juvenile striped bass were de- fi-osted, and standard, fork, and total lengths (SL, FL, and TL respectively) were measured. Sagittal otoliths were removed from each fish, cleaned in 10% sodium hypochlorite (bleach), and embedded in Spurr's epoxy. One otolith from each pair was subse- quently sectioned along the transverse axis and pre- pared in thin section according to the method of Secor et al. (1991). A total of 63 juvenile otoliths were suc- cessfully prepared and read. Otolith microstructure analysis Otolith microstructure was examined with a light microscope with a video attachment connected to an image analysis system. Specimens were viewed in the sagittal plane by using the digital image for pro- jection. Most specimens were viewed at magnifica- tions of 500-750X, but 1250x (lOOx objective) was used to resolve the finest increments. Increments were counted several times and averaged. Larval otoliths were examined by one reader, and two inde- pendent readers were used to cross-check the con- sistency of increment counts for juvenile striped bass otoliths. Ages were estimated by correcting for the effect of temperature on the timing of first increment deposi- tion (Houde and Morin^ ). Age at first increment depo- sition was estimated as (11.56-0.45 T) for striped bass and (9.05-0.327) for white perch, where T is water temperature (Houde and MorinM. From our measurements of water temperature («=30) from 14 April to 19 July 1994, we estimated a linear increase in river temperatures over the study period with the regression Water temperature - -17.6 -t- 0.23 x (day of year) [r2=0.98, P<10-*1. We used this equation to estimate water tempera- tures on the day of the first increment and assumed little change in temperature back to the hatching day. Ages were estimated as number of increments + tem- perature adjustment; hatching dates were then backcalculated from the date of capture. Image analysis was used to measure increment widths. For larvae, widths of the total number of in- crements on an otolith were measured and averaged to obtain an average daily increment width (in mm). In addition, when possible, widths of the first five increments (11-15) and of the sixth and seventh in- crements (16-17) were measured to obtain two stan- dardized estimates of early growth. Increments Il- ls exhibited slower average growth than did incre- ments 16-17. In all, otoliths of 248 larval striped bass and 526 larval white perch were examined. Somatic growth rates were calculated for larvae and juveniles by the equation ^^,-K )IAge, where L, = length at capture; and L, = length at hatch. L^ is assumed to be 3.0 mm for white perch and 4.0 mm for striped bass (Mansueti, 1958; Lippson and Moran, 1974). In addition, "standardized" growth rates were calculated for 7-day-old larvae so that comparisons of growth could be made before, dur- ing, and after the zooplankton bloom. This was done by determining the number of increments deposited by age 7 (typically <5), multiplying that number by the mean increment width of 11-15, adding the re- sult to the otolith core radius, and then by using the following regressions to calculate standard lengths at age 7 days: SLgg = -0.37 -I- 1.95 (natural log of otolith width) [r2=0.83, «=210, P<10-*], SL UP ■ -1.67 + 2.18 (natural log of otolith width) [/-'■^=0.88, /!=522, P<10-*]. Houde, E. D., and L. G. Morin. 1990. Temperature effects on otolith daily increment deposition in striped bass and white perch larvae. International Council for the Exploration of the Sea, Copenhagen, Denmark, Council Meeting 1990/M:5. The same procedure was used to backcalculate age 7-d somatic growth rates for juvenile striped bass. Mortality and recruitment potential For purposes of comparison with Chesapeake Bay studies, mortality rates were estimated for fish lar- vae collected at New Hamburg (rkm 105) by using a method similar to that described in Secor and Houde (1995). Cohort analyses were confined to the New Hamburg station because we were able to collect both species consistently in large numbers. All larvae in these samples were counted, and lengths of at least Limburg et al.: Growth, mortality, and recruitment of larval Morone spp. 83 100 individuals were measured unless fewer than 100 occurred in the sample. Densities (numbers of fish per 1000 m^) by 0.5-mm size class were com- puted and corrected for differences in day and night catch efficiencies, and for extrusion of smallest indi- viduals, by using regressions in Houde et al.^ Age distributions offish collected on different sam- pling dates were estimated from the otolith-aged fish. Rather than estimate variance from an age-length regression and use it to assign probabilities of age- at-length (Secor and Houde, 1995), we estimated the mean and variance in ages for each 0.5-mm size group and used those parameters to assign age-at-length probabilities (z-scores). The corrected catch densities, separated into 0.5- mm size classes, were multiplied by the appropriate size-based age distributions to obtain numbers offish of different ages. Following Secor and Houde (1995), hatching dates were calculated and fish were grouped into 6-d cohorts. In some of the cohorts, the age-esti- mation technique resulted in very small initial num- bers of fish. If these numbers were tenfold smaller than the maximum calculated densities of a cohort, they were deleted from the data set. Then mortality rates (Z) of the cohorts were estimated by fitting an exponential decay model to abundances offish within specific cohorts over time. We used an index (Rutherford and Houde, 1995; Secor and Houde, 1995) of mean instantaneous growth (estimated for an individual fish as G = In (W/Wq)//, where t = age at capture, W, = weight at capture, and W^ = weight at hatch) over instanta- neous mortality rate iG/Z) to compare the benefits accrued in growth versus the costs expressed as mor- tality for larvae occurring before, during, and after the zooplankton bloom. Weights were estimated from lengths by the equation Table 1 Mean (± SD) somatic growth rates (GR, mm/d) of larval white perch and striped bass, by site and date, 1994. (n=526 white perch and 248 striped bass; n.d.=no data; — = zero fish in samples.) White perch Striped bass Date MeanGR SD n Mean GR SD Haverstraw Bay (rkm 65-70) 18 May 24 May 3 June 7 June 10 June 13 June 22 June 29 June 6 July 0.181 0.067 0.186 0.409 0.211 0.382 0.427 0.313 n.d. 0.054 0.215 0.086 0.199 New Hamburg (rkm 105) 18 May 24 May 3 June 7 June 10 June 13 June 22 June 29 June 6 July 0.155 0.192 0.218 0.236 0.210 0.265 0.306 0.331 Kingston (rkm 148) 18 May — 24 May 0.205 3 June 0.167 7 June 0.200 10 June 0.201 13 June 0.196 22 June 0.245 29 June 0.300 6 July n.d. 0.037 0.060 0.070 0.085 0.081 0.083 0.049 0.052 0.061 0.053 0.053 0.089 0.086 0.113 0.099 15 0 17 9 8 1 3 1 0 37 47 42 35 29 11 11 9 0 9 52 47 59 35 12 37 0.072 0.069 0.185 0.286 0.178 0.250 0.293 0.284 n.d. 0.069 0.162 0.216 0.247 0.224 0.207 0.252 0.264 0.044 0.032 0.146 0.017 n.d. 0.111 0.128 0.111 0.046 0.088 0.007 0.022 0.127 0.084 0.030 0.039 0.056 0.084 0.062 0.027 0.054 0.189 9 0 29 25 8 17 36 3 0 3 11 14 4 8 9 20 40 0 3 6 2 1 0 0 0 W = 3.763 X 10-1 ^ ^(4 2879) [n=67, r2=0.92, P<10-6]; where W = mg wet weight; and L - length in mm (Limburg et al. 1997). Finally, we estimated the dates of first-feeding for striped bass larvae and juveniles with respect to the zooplankton bloom. We assumed that fish begin to feed at day 5 after hatching. We used an age-length regression from the 63 juvenile striped bass to esti- mate ages of the remaining juveniles. We assumed that little or no out-migration would be occurring in July and August (Dovel, 1992), so that fish collected during these months would show the effect of differ- ential recruitment without the confounding process of out-migration. Results 2 Houde, E. D., E. J. Chesney R. M. Nyman, and E. S. Ruther- ford. 1988. Mortality, growth and growth rate variability of striped bass larvae in Chesapeake subestuaries. Interim Re- port to Maryland Department of Natural Resources. Chesapeake Biological Laboratory, Solomons, MD. Ref No. [UMCEES]CBL 88-96, 126 p. [Available: Chesapeake Biological Laboratory, Box 38, Solomons, MD 20688.) Growth rates Mean somatic growth rates (GRs) of both species of Morone tended to be lowest prior to the bloom (be- fore 3 June); rates were higher after the bloom (22 June and later), especially for white perch (Table 1; 84 Fishery Bulletin 97(1), 1999 Fig. 1). There were significant site, species, and site X species effects on mean somatic GRs (P<10~'*), with highest GRs at the southernmost site (Haverstraw Bay). Mean somatic growth rate for striped bass was 0.053, 0.232, and 0.238 mm/d at Kingston, New Ham- burg, and Haverstraw Bay, respectively (greater than an eightfold difference between highest and lowest; it should be noted that the total number of fish sampled at Kingston was very small). Growth rates for white perch did not vary as greatly but did vary in the same longitudinal direction (0.209, 0.215, and 0.245 mm/d at Kingston, New Hamburg, and Haverstraw Bay). Time series of somatic GRs, aver- aged over all sites for each species (Fig. 1), revealed species-specific patterns of growth with respect to the zooplankton bloom period. Striped bass GRs ac- celerated more strongly during the bloom period than did GRs of white perch. Somatic growth rates were related to water tem- perature as GR = -0.050 -I- 0.013 X MWT [/•2=0.15, /i=774, P<10-*^], where MWT = mean water temperature during the life of the individual larva (calculated as the tem- perature midway through the larva's life). The effect of temperature was removed by calculat- ing residuals of this regression and by using the re- 0.40 0.35 0.30 ( k 0.25 >, ■^ 020 i 1 . ■ •♦ i a E 0 15 E 0 0 \ + 0.10 white perch striped bass 0.05 0 00 18-May 28-May 7-Jun 17-Jun 27-Jun 7-Jul 18-May 28-May 7-Jun 17-Jun 27-Jun 7-Jul Date Figure 1 Mean (±SE) somatic growth rates of larval white perch and .striped bass, averaged across all sampling sites, 18 May-6 July 1994. The darkened points on each graph indicate the period of the Bosmina bloom. siduals as a new data set ( temperature-adjusted GRs) to assess growth relative to zooplankton dynamics. Temperature-adjusted somatic GRs for white perch showed little trend with respect to the zooplankton bloom, whereas striped bass GRs peaked during the bloom ( Fig. 2 ). This peak was not significantly greater than the adjusted GRs after the bloom, but both dif- fered significantly from the prebloom-adjusted GRs. Our comparisons of somatic growth rates are based on fish larvae of different sizes and ages. Compari- sons of growth rates for a standardized age (we chose age=7 days) can eliminate this potential source of bias. Mean 7-d GRs (Fig. 3) showed a pattern simi- lar to that for overall mean GR, indicating that growth rates did not peak during the zooplankton bloom. Although both species' 7-d GRs varied signifi- cantly with respect to the bloom (with highest GRs after the bloom), white perch 7-d GRs did not differ significantly from striped bass 7-d GRs. Recruitment potential at New Hamburg Ten 6-d age cohorts, grouped arbitrarilv starting 1 May (cohort A) and extending through the period beginning 24 June (cohort J), were identified in tin; catches at New Hamburg (rkm 105). Insufficient numbers prohibited estimation of Z for the last striped bass cohort (J, 24-29 June) and the last two white perch cohorts (I and J). Estimates of white perch starting densities (natural log of A^,,) were high- est in early cohorts (A and B; Table 2). Striped bass initial density estimates were highest in later cohorts (E, G, and H; Table 2). Earliest and latest cohorts were poorly estimated owing to low sample numbers. Estimated mortality rates were highest in cohort A for both species (Table 2). Cohort A (period starting 1 May) ap- peared only in two samples; presumably recruitment from this group was very low. For white perch, mortality rates declined more or less mono- tonically with time (and there- fore also with increasing tem- perature). Striped bass mortal- ity rates were also lowest in the latest cohorts, but there did not appear to be a system- atic decline (Table 2). Instantaneous growth rates (G) in striped bass did not vary Limburg et al.: Growth, mortality, and recruitment of larval Morone spp. 85 JC o 0.06 0.02 -0.02 -0.06 -0.10 -0.14 striped bass white perch Before During After Before During After Zooplankton bloom status Figure 2 Mean (±SE) somatic growth rates (residuals) of larval white perch and striped bass with the effect of tem- perature removed, grouped with respect to the Bosmina bloom. 0.40 0.10 T3 E E 0.35 striped bass ra 1 0.30 T3 0) to o 0.25 0.20 ^ % o (5 0,15 'Y white perch Before During After Before During After Zooplankton bloom status Figure 3 Mean (iSE) growth rates of 7-day old larval white perch and striped bass, grouped with respect to the Bosmina bloom. in a consistent manner, either with temperature or zooplankton density (Table 3). Striped bass G peaked in cohort E (those hatched 25-30 May), as did G/Z. Neither G, Z, nor G/Z for striped bass were significantly affected by either temperature or zooplankton densi- ties estimated during the week of hatching. 86 Fishery Bulletin 97(1), 1999 Table 2 Estimates of natural logs of initial densities (A^^. lumbers per 1000 m^) and mortality rates (Zl for 6-d cohorts of striped bass and white perch in the Hudson River (New Hamburg site). Letters designate cohorts referred to in text. Date Cohort ln(Ar„) SE Z SE r2 P n Striped bass 1-6 May A 4.132 1.51 2 7-12 May B 7.598 6.153 0.262 0.315 0.41 0.55 3 13-18 May C 6.919 2.070 0.263 0.132 0.66 0.19 4 18-24 May D 5.941 2.207 0.296 0.110 0.64 0.05 6 25-30 May E 7.690 0.751 0.260 0.034 0.94 0.001 6 31 May-5 June F 6.502 0.549 0.176 0.027 0.91 0.002 6 6-11 June G 8.808 0.103 0.221 0.005 0.99 0.01 3 12-17 June H 7.645 0.767 0.163 0.048 0.92 0.18 3 18-23 June I 6.977 0.145 2 White perch 1-6 May A 16.653 0.785 2 7-12 May B 12.484 2.905 0.509 0.138 0.87 0.07 4 13-18 May C 9.221 1.299 0.328 0.064 0.87 0.007 6 18-24 May D 8.058 1.284 0.241 0.064 0.78 0.02 6 25-30 May E 9.682 1.022 0.304 0.051 0.90 0.004 6 31 May-5 June F 7.861 0.807 0.187 0.049 0.83 0.03 5 6-11 June G 5.811 1.295 0.044 0.087 0.21 0.70 3 12-17 June H 7.885 0.137 2 18-23 June I - insufficient data Table 3 Estima ted instantaneous growth (G) rates (± standard error), and index of recruitment potenti al iGlZ) for 6-d cohorts of larval striped bass ( SB) and white perch (WP) New Hamburg (rkm 105), 1994 . Calculated water te mperatures (see "Methods") and zooplan kton densities (copepods and Bosmina no./L), correspo iding to eac h cohort are included Zooplankton densities are offset by 5-7 d from actual sample dates (e.g. the zooplankton sample associated with cohort G was collected on 10 June ) to approximate the lag in initial feeding by 1 arvae. n.d. = no data. G/Z in parer theses were calculated with the previous cohorts' Z estimates but | were not used in analyses. Temperature Zooplankton Cohorts Dates (°C) density G'sB SE G/ZsB <5wp SE GIZ^ A 30 Apr 1994 B 5 May 1994 12.2 n.d. 0.125 0.006 0.245 C 12 May 1994 13.6 1.88 0.205 0.056 0.779 0.125 0.006 0.380 D 18 May 1994 15.0 10.9 0.205 0.056 0.693 0.141 0.043 0.583 E 24 May 1994 16.4 35 0.564 — 2.169 0.132 0.010 0.436 F 30 May 1994 18.0 80.5 0.070 0.009 0.398 0.159 0.007 0.850 G 05 Jun 1994 19.4 520 0.092 0.012 0.416 0.378 — 8.521 H 11 Jun 1994 20.8 70 0.113 0.021 0.693 0.196 — 1.427 1 17 Jun 1994 22.2 4.1 0.097 0.035 0.671 0.150 0.036 (1.095) J 23 Jun 1994 23.5 5.6 0.171 0.228 (1.179) In contrast, white perch G was significantly corre- lated with zooplankton density (G\^=0. 136+4.64x10"^ xizooplankton density), r''^=0.96, P<10 ''). The high significance was due to the highest growth at a time of maximum zooplankton density (520 individuals/ L), but the regression relationship remained the same even with removal of the infiuential point. Lowest mean G/Z occurred before the bloom (0.35 ±0.18 SD, M=4), intermediate G/Z during the bloom (1.27 ±1.06, n=2) and highest afterward (2.53 ±2.04, n=2), but these were not significantly different. Although there is a tendency toward increasing G/Z with time (and temperature), it was not statistically significant either. Limburg et al.: Growth, mortality, and recruitment of larval Morone spp. 87 Table 4 Mean growth (mm/d, calculated from otoliths) at age 7 d after hatching of larval and juvenile striped bass, 1994. Bloom status Mean SD n Larvae Before 0.167 0.072 45 During 0.221 0.062 47 After 0.331 0.079 77 Juveniles Before 0.199 0.080 18 During 0.215 0.063 33 After 0.223 0.056 12 Retrospective early life histories of juvenile versus larval striped bass Larval and juvenile striped bass from the 1994 year class had similar patterns of early growth, as mea- sured by mean growth (mm/d I through age 7 days after hatching. In both groups, growth rates in- creased monotonically from May through early July corresponding with before, during, and after the zoo- plankton bloom (P<0.01; Table 4). Juvenile mean growth rates before and during the zooplankton bloom were higher (though not significantly so) than larval growth rates in the corresponding periods. However, mean 7-d growth rates of postbloom juve- niles were significantly {F( 1,87)=21.03, P<10-*) lower than growth rates of postbloom larvae. The frequency (distribution of hatching dates) of juvenile striped bass born in the 6-d cohorts begin- ning 1 May did not correlate well with the index of recruitment (G/Z) developed from the larval data {Frequency=0.n8+0.021 iG/Z), r^=0.03). Because of our ability to "hind-cast" the distribu- tion of hatching dates of juveniles and larvae, we estimated the relative proportions of both larvae and juvenile striped bass that would have begun to feed during the zooplankton bloom (assuming that feed- ing begins at day 5 after hatching). For this analy- sis, we analyzed the fish response to Bosmina and copepods. For Bosmina we assumed that fish with hatching dates between 29 May and 8 June would be able to feed on Bosmina: we further assumed that fish hatched between 5 June and 17 June would also be able to take advantage of the bloom of copepods (note that in this analysis we assume that Bosmina- feeders occurred above Haverstraw Bay and that copepod feeders occurred throughout the river). Larval distributions indicated that 3T7( of the popu- lation began to feed after the bloom of Bosmina (Fig. 4A), and 26% initiated feeding during the bloom. Of larvae that occurred in the geographic range of Bosmina (at New Hamburg and Kingston), over half began to feed after the bloom and less than 20% began feeding during the bloom. In contrast, 44% of juvenile striped Pre During I Larvae □Juveniles Post Pre During [Larvae □Juveniles Post Figure 4 Fraction of larval (black bars) and juvenile (gray bars) 1994 striped bass populations that were estimated to be feeding before, during, and after blooms of (A) Bosmina freyi and (B) copepods in the Hudson River. bass began to feed during the bloom of Bosmina (Fig. 4A); and of the fish identified as first feeding prior to the onset of the bloom, 23% were hatched after 25 May, so that these fish would have been relatively close to first feeding at the time that Bosmina numbers in- creased exponentially. Those survivors to the juvenile stage that began to feed after the Bosmina bloom com- posed the smallest fraction ( 26% ). The first-feeding dis- tributions of larvae and juveniles differed significantly with respect to the Bosmina bloom (2x3 contingency analysis, x^=13.4, df=2, P<0.01 ), whereas distributions of larvae and juvenile fish with respect to the bloom of copepods did not differ from one another (x'=3.7; Fig. 4B). Thus the time period associated with the Bosmina bloom, and not the copepod bloom, appears to have been related to successful striped bass recruitment. Discussion Evidence from this study provides some support for the relation between larval growth rates and tern- 88 Fishery Bulletin 97(1), 1999 perature found by Rutherford and Houde ( 1995) for the Potomac River and Upper Chesapeake Bay and by Secor and Houde ( 1995) for striped bass larvae in the Patuxent River. However, the relationship we found was weak (/•" only 0.15). Some measures of growth were also enhanced with increased food avail- ability. Because of the similarity of our approach with that of these previous Chesapeake researchers, we can compare the fates of striped bass larvae in the Chesapeake and Hudson systems. Temperatures during the larval period were more variable in the Chesapeake tributaries in the stud- ies referred to above than in the Hudson in 1994. In the Patuxent in 1991, water temperatures in April (when nearly 90'^ of eggs were produced) fluctuated erratically between approximately 13° and 20°C. In May, temperatures rose through the month from 20° to 29°C (Secor and Houde, 1995). In the Potomac (1987-89) and Upper Bay ( 1988-89) (Rutherford and Houde, 1995), water temperatures also fluctuated, dropping in 1989 during one event from above 16"C to around 12°C, the lower lethal limit for striped bass larvae (Morgan et al., 1981). April and May water temperatures in the Nanticoke, another Chesapeake tributary, also fluctuated in 1992 and 1993 (Kellogg et al.^). In contrast, 1994 water temperatures in the Hudson River increased linearly from mid-April through mid-July with little fluctuation. In spite of the overall warmer temperatures, cal- culated instantaneous growth rates of larval striped bass grouped into 6-d cohorts were lower in the Patuxent River (mean instantaneous G=0.126/d. range 0.1 1-0.14) than in the Hudson (mean G=0.190/ d, range 0.070-0.564). Nanticoke River growth rates were intermediate (0.166 in 1992, 0.159 in 1993; Kellogg et al.'^). Hudson striped bass larvae grew at rates similar to those for Potomac River (0.208 and 0.181/d for 1987 and 1989) and Upper Chesapeake Bay larvae (0.183/d). In terms of growth in length, both moronid species in the Hudson grew faster than Patuxent River striped bass larvae (Hudson white perch=0.212 mm/d ±0.101 SD, Hudson striped bass=0.218 mm/d ±0.121 SD, Patuxent striped bass=0.17 mm/d for 0-25 d). Note that Dey (1981) estimated Hudson River striped bass larval growth rates (in 1973-76) as ranging from 0.1 to 0.2 mm/d. ^ Kellogg, L L., E. D. Houde, and D. H. Secor. 1996. Egg pro- duction and environmental factors influencing larval popula- tion dynamics in tlie Nanticoke River, 1992-1993. Chapter 1 in E. D. Houde and U. H. Secor (eds.). Episodic water quality events and striped bass recruitment: lar\',al mark-recapture experi- ments in the Nanticoke River Final Rep. to Maryland Dep. Natural Resources, Chesapeake Bay Research and Monitoring Division. University of Maryland, Ref No. lUMCEESICBL 9(v 083. [Available: Chesapeake Biological Laboratory, Box 38, Solomons, MD 20688.! based on analysis of weekly changes in mean lengths of larvae. Growth rates in the present study are higher than Day's estimate and fall within the range of the Potomac (mean growth rates 0.18-0.26 mm/d, 1987-89; Rutherford and Houde, 1995). Differences in growth rates are likely due to a combination of factors in any given year and site; unfortunately, multiyear studies of larval growth rates, which would provide ranges of variation, are few. An index of population growth, Gl Z, has been used as an index of striped bass recruitment success (Ru- therford and Houde, 1995, Secor and Houde, 1995, Rutherford et al., 1997, Kellogg et al.^) in Chesapeake tributaries because the index corresponded, at least on a seasonal basis, with production of 8-mm larvae, which is, in turn, correlated with year-class strength in the Chesapeake system. The GIZ did not corre- late well with our index of recruitment, i.e. juvenile striped bass recruits. We did not examine juvenile recruitment of white perch, nor are we aware of any otolith-based studies of white perch recruitment in other ecosystems. The GIZ method as applied at New Hamburg must be viewed cautiously, as calculations of mortality (Z) for individual cohorts are based on a few samples and are not well constrained (hence the large stan- dard errors for some cohorts, Table 2). Further, al- though in aggregate a large number of individuals were examined in this study, growth rates for indi- vidual cohorts were estimated from a limited sample size. Finally, there is likely a substantial variability in time and space for both G and Z that currently exceeds measurement capacity even in the most dili- gent field study. Given these issues of accuracy and precision, our analyses should indicate general trends, but they cannot yet be viewed as conclusive. In both the Chesapeake Bay and Hudson River, it appears that lowest mortality rates in striped bass larvae fall within a constrained temperature range. Secor and Houde (1995) found a convex parabolic relationship of Patuxent striped bass mortality rates with temperature; minimum mortality rates were associated with cohorts that experienced water tem- peratures of 16-18°C in the first 25 days of life. Ru- therford and Houde ( 1995) did not find so neat a re- lationship but rather noted that low mortality rates occurred in early May cohorts of Potomac striped bass larvae, which coincided with water temperatures around 16°C. Mortality rates of Hudson River striped bass in 1994 were also lowest in cohorts associated with water temperatures of 16.4° to 18.0°C during the week of hatching (31 May to 5 June, Table 2); however, first feeding by these cohorts also coincided with the zooplankton blooms. White perch mortality rates appeared to decline with increasing temperature. Limburg et al : Growth, mortality, and recruitment of larval Morone spp 89 In the Chesapeake system, temperature can play a key role in setting year-class strength for striped bass (Rutherford and Houde, 1995; Secor and Houde, 1995), although it is not al- ways the determining factor (e.g. Kellogg et al."^). Complete failures of individual cohorts were associated with rapid temperature drops down to 12°C (Rutherford and Houde, 1995). Dey (1981) hypothesized that early cohorts of Hudson River striped bass were eliminated by a sudden drop in water temperatures in late May 1976, and simulation analysis by Boreman (1983) supported this inference. Nevertheless, there was no relationship be- tween temperature and year-class strength for Hudson River moronids over the period 1974- 90 (Pace et al., 1993). Examining a 40-yr record of Hudson River temperatures, we found that the likelihood of lethal low temperature events (defined as <12°C) declines rapidly during the latter half of April (Fig. 5). Early spawned co- horts of both moronid species have a risk for low-temperature mortality events, but most co- horts are spawned after early May and are highly unlikely to experience this direct source of mortality (Fig. 5). We have demonstrated that, like Chesapeake striped bass growth rates, larval moronid growth rates are linked (albeit weakly) to tem- perature in the Hudson River Nevertheless, differential mortality occurred on the faster- growing, late cohorts of striped bass. We postu- late that this mortality was due to predation. Secor and Houde (1995) also speculated that increased mortality rates observed in late Patuxent cohorts was due to predation. Hudson River moronid larvae that co-occurred with the zooplankton blooms were able to benefit ener- getically from increased food availability in re- lation to prebloom conditions (Limburg et al., 1997). Consumption rates continued to be high after the bloom, however, so that energetically the larvae continued to do well. Warmer temperatures after the bloom would have enabled larvae to swim faster and encounter more prey, but they would also potentially have encountered more predators. Thus, we infer that late cohorts suffered differentially greater predation. We suggest that larger estuarine nurseries, such as the Hudson River, provide more damping of physi- cal fluctuations in characteristics such as water tem- perature than do smaller estuaries with "flashier" drainage regimes. If this is the case, we might ex- pect Chesapeake Bay moronid stocks to be more at risk for episodic, density-independent mortality events of the type described in Rutherford and Houde 14 12 10 8 6 4 2 0 Thermal Risk- -D- SB. eggs -O- SB.ysl -A- SB.pysl \ b ^ U-U-U-U'U 1.0 0.8 06 0.4 0.2 0.0 Maris Apr 7 Apr 27 May 17 Jun 6 Jun 26 Jul 16 Aug 5 a c 14 12-1 10 8- 6- 4 2 0-1 -2- Maria Pj.Q ThermalRisk- □ 'b-&'0;A-A-A^ -D-WP.eggs /O / O-S^ A _o_vVP.ysl / A / ^ ''• p o ;^-A- WPpysl 1.0 0.8 0.6 04 0.2 00 o a (D 3 ro o o Apr 7 Apr 27 May 17 Jun 6 Jun 26 Jul 16 Aug 5 Week Figure 5 Probability of water temperature occurring at or below 12°C in Hudson River (based on 40-year database: Poughkeepsie Water Works. Poughkeepsie, NY) compared with means of log-trans- formed, river-wide abundances (1000s) of early life stages of (A) striped bass and (B) white perch. Squares = eggs, circles = yolksac larvae, and triangles = postyolksac larvae. (1995). Klauda et al. (1980) noted that striped bass year-class strength, indexed by juvenile abundance, varied about 34-fold in the Hudson River, about 47- fold in Chesapeake Bay, and 160-fold in the Roanoke River (where upstream water temperatures are con- trolled by water regulation). The different lengths of the data sets, as well as the fact that Chesapeake Bay juvenile index reflects a mix offish derived from tributaries and the Bay proper, make this compari- son somewhat problematic to interpret. What is likely is that increased physical variability in smaller estu- aries translate into greater recruitment variability. Further, evidence from smaller streams suggests that recruitment improves in years with greater spring runoff (McGovern and Olney, 1996; Kellogg 90 Fishery Bulletin 97(1), 1999 et al.'^). Increased runoff in small streams could trans- late into more available habitat for larvae, including greater food availability (Bulak et al. 1997; Kellogg et al.'*). Larger watercourses like the Hudson should, on average, provide a more constant amount of suit- able habitat between years. Further, we suggest that biotic controls on recruit- ment, if not as easily assessed as temperature and discharge, may play relatively greater roles in larger estuaries than in smaller ones, or at least in those systems with relatively predictable physical forcing. Our results implicate biotic factors for one year class of striped bass in the Hudson and are based on a comprehensive approach that included careful obser- vation of larvae and retrospective analysis of juve- nile early life histories. Had we not carried out the retrospective analysis, we likely would have pre- dicted that later cohorts would have had the great- est recruitment potential. Acknowledgments We thank K. Hattala, A. Kahnle, and K. McKown, New York Department of Environmental Conserva- tion, for providing juvenile striped bass. E. Houde and D. Secor, Chesapeake Biological Laboratory, and E. Rutherford, University of Michigan, provided help- ful advice on otoliths and general analysis. Three anonymous reviewers gave constructive comments. Support for this work was provided by the Hudson River Foundation, New York Sea Grant, and by a Tibor T. Polgar fellowship. Literature cited Boreman, J. 1983. Simulation of striped ba.ss egg and larva development based on temperature. Trans. Am. Fish. Soc. 1 12:286-292. Boreman, J., and R. J. Klauda. 1988. Distributions of early life stages of striped bass in the Hudson River estuary, 1974-1979. Am. Fish. Soc. Monogr. 4:5.3-.58. Brothers, E. B. 1987. Methodological approaches to the examination of otoliths in aging studies. In R. C. Summerfelt and G. E. Hall (eds.), Age and growth of fishes, p. 319-3-30. Iowa State Univ. Press, Ames, lA. Bulak, J. S., J. S. Crane, D. H. Secor, and J. M. Dean. 1997. Recruitment dynamics of striped bass in the Santee- Cooper system, South Carolina. Trans. Am Fish. Soc. 126:133-143. Campana, S. E., and J. D. Neilson. 1985. Microstructure of fish otoliths. Can. J. Fish. Aquat. Sci. 42:1014-1032. Chesney, E. J., Jr. 1989. Estimating the food requirements of striped bass lar- vae Morone saxatilis: effects of light, turbidity and turbulence. Mar Ecol. Prog. Ser. .53:191-200. Cowan, J. H, Jr., K. A. Rose, E. S. Rutherford, and E. D. Houde. 1993. Individual-based model of young-of-the-year striped bass population dynamics II. Factors affecting recruitment in the Potomac River, Maryland. Trans. Am. Fish. Soc. 122:439-458. Dey, W. P. 1981. Mortality and growth of young-of-the-year striped bass in the Hudson River Estuary. Trans. Am. Fish. Soc. 110:151-157. Dovel, W. L. 1992. Movements of immature striped bass in the Hudson estuary. In C. L. Smith (ed.), Estuarine research in the 1980s, p 276-300, SUNY Press. Albany, NY. Hilborn, R., and C. J. Walters. 1992. Quantitative fisheries stock assessment: choice, dy- namics, and uncertainty. Chapman and Hall, New York, NY, 570 p. Klauda, R. J., W. P. Dey, T. B. Hoff, J. B. McLaren, and Q. E. Ross. 1980. Biology of Hudson River juvenile striped bass. In H. Clepper(ed.), Proceedings of the 5"' annual marine rec- reation fishery symposium, p. 101-123. Sport Fishing Institute, Washington. D. C, Klauda, R. J., J. B. McLaren, R. E. Schmidt, and W. P. Dey. 1988. Life history of white perch in the Hudson River estuary. Am. Fish. Soc. Monogr. 4:69-88. Leggett, W. C, and E. DeBIois. 1994. Recruitment in marine fishes: Is it regulated by star- vation and predation in the egg and larval stages? Neth. J. Sea Res. 32:119-134. Limburg, K. E., M. L. Face, D. Fischer, and K. K. Arend. 1997. Consumption, selectivity, and utilization of zooplank- ton by larval Morone spp. in a seasonally pulsed estuary. Trans. Am. Fish. Soc. 125:607-621. Lippson, A. J., and R. L. Moran. 1974. Manual for identification of early developmental stages of fishes of the Potomac estuary. Prepared for MD Power Plant Siting Program, PPSP-MP-13. Martin Marietta Corp., Baltimore, MD. 282 p. Mansueti, R. 1958. Eggs, larvae and young of the striped bass, Roccus saxatilis. Maryland Dep. Res. Educat. Contrib. 112, 35 p. Margulies, D. 1989. Effects of food concentration and temperature on development, growth, and survival of white perch. Morone amcricann. eggs and larvae. Fish. Bull. 87:63-72. McGovern, J. C, and J. E. Olney. 1996. Factors affecting survival of early life stages and subsequent recruitment of striped bass on the Pamunkey River. Virginia. Can. J. Fish. Aquat. Sci. 53:1713-1726. Morgan, R. P., Ill, V. J. Rasin, and R. L. Copp. 1981. Temperature and salinity effects on development of striped bass eggs and larvae. Trans. Am, Fish. Soc. 110:9.5-99. Pace, M. L., S. B. Baines, H. Cyr, and J. A. Downing. 1993. Relationships among early life stages of Morone americana and Morone saxatilis from long-term monitor- ing of the Hudson River estuary. Can. J. Fish, Aquat. Sci. .50:1976-1985. Rogers, B. A., and D. T. Westin. 1981. Laboratory studies on effects of temperature and delayed initial feeding on development of striped bass larvae. Trans. Am. Fish. Soc. 110:100-110. Limburg et al.: Growth, mortality, and recruitment of larval Morone spp. 91 Rutherford, E. S. 1992. Relationship of larval stage growth and mortality to recruitment of striped bass, Morone saxatilis, in Chesa- peake Bay. Ph.D. diss., Univ. Maryland, College Park, MD, 369 p. Rutherford, E. S., and E. D. Houde. 1995. The influence of temperature on cohort-specific growth, survival, and recruitment of striped bass, Morone saxatilis, larvae in Chesapeake Bay. Fish. Bull. 93:315- 332. Rutherford, E. S., E. D. Houde, and R. M. Nyman. 1997. Relationship of larval-stage growth and mortality to recruitment of striped bass, Morone saxatilis. in Ches- apeake Bay. Estuaries 20:174-198. Secor, D. H., J. M. Dean, and S. E. Campana. 1995. Recent developments in fish otolith research. Univ. South Carolina Press, Columbia, SC, 735 p. Secor, D. H., J. M. Dean, and E. H. Laban. 1991. Manual for otolith removal and preparation for mi- crostructure examination. Belle W. Baruch Institute for Marine Biological and Coastal Research. Univ. South Carolina, Columbia, SC, 85 p. Secor, D. H. and E. D. Houde. 1995. Temperature effects on the timing of striped bass egg production, larval viability, and recruitment potential in the Patuxent River (Chesapeake Bay). Estuaries 18:527- 544. Tsai, C.-F. 1991. Prey density requirements of the striped bass, Morone saxatilis (Walbaum) larvae. Estuaries 14:207-217. Uphoff, J. H., Jr. 1989. Environmental effects on survival of eggs, larvae, and juveniles of striped bass in the Choptank River, Maryland. Trans. Am. Fish. Soc. 118:251-263. 92 Abstract— The roudi escolar Pro- methichthys prometheus is common in deep hook-and-line and longline catches of a small-scale fishery along the slope off the Canary Islands. Population structure, reproduction, growth, and mortality of the species were studied from sampling undertaken from August 1992 to July 1995. Range of length of fish in the catches was between 36 and 80 cm TL, with a main distribution be- tween 56 and 66 cm. The overall ratio ofmales to females was 1:1.74. Females predominated in all sizes. The sex ra- tio varied throughout the period of study; the lowest discrepancy between males and females, however, was dur- ing the reproductive period. A vertical space partitioning among sexes was ob- ser\'ed. with males predominating from 600 to 800 m depth, females from 300 to 500 m. The reproductive period of the species was from April to September, with a peak in spawning in June-July. The size at first maturity was 47.41 cm. The parameters of the length-weight relationship for all fish were 0=0.004521 and 6=2.98932. Age read- ings of otoliths indicated that the ex- ploited population consisted of nine age groups (III-XI years). The von Bertalanffy growth parameters for all individuals were L^=93.61 cm, ^=0.18/ years, and /„ =-1.54 years. The rates of mortality for all fish were Z=0.49/years, M=0.3.5/years, and F=0.14/years. The length at first capture for the whole population was 51.57 cm. Biology of a deep benthopelagic fish, roudi escolar Promethichthys prometheus (Gempylidae), off the Canary islands Jose M. Lorenzo Jose G. Pajuelo Departamento de Biologia (Universidad de Las Palmas de Gran Canana) Edificio de Ciencias Basicas, Campus Universitano de Tafira 35017 Las Palmas, Spam E-mail address (for J. M, Lorenzo) losemaria lorenzojSbiologia ulpgc es Manuscript accepted 23 April 1998. Fish. Bull 97:92-99 (1999). The family Gempylidae consists of 16 genera and 23 species. Only seven species are found off the Ca- nary Islands, one of which is the roudi escolar Promethichthys pro- metheus (Cuvier, 1832), the only species recognized to date in the genus Promethichthys (Nakamura and Parin, 1993). The roudi escolar is a bentho- pelagic marine fish that has a worldwide distribution in tropical and warm temperate waters. This species generally inhabits waters between 100 and 800 m in depth over seamounts and continental and insular slopes. It migrates up- ward at night, probably forming schools (Nakamura, 1981; Parin, 1986; Nakamura and Parin, 1993). Published information on P. prometheus is very scarce. The ma- jority of studies describe its morpho- logical characteristics, geographical and depth distribution, and ecology (Nakamura, 1981; Parin, 1986; Nishikawa, 1987; Nakamura and Parin, 1993). Only Lorenzo and Pajuelo (1995) have studied some biological aspects of the species. These authors carried out a prelimi- nary study on the sex ratio, repro- duction, and age and growth of roudi escolar off the Canary Islands (central-east Atlantic) on the basis of a small number of specimens dur- ing one life cycle. This paper is an extension of their work, analyzing, in addition to all those aspects, population structure and mortality. The roudi escolar is common in the catches of the deep hook-and- line and longline small-scale fish- ery over the slope off the Canary Islands. In this area, this species is captured year round without signifi- cant seasonal differences in landings. Materials and methods Between August 1992 and July 1995, the TL (cm) of 1879 specimens of roudi escolar was measured monthly from commercial catches of the small-scale fleet. Fish were caught with baited hook-and-lines and longlines at depths of 285-870 m around the islands of the Canary archipelago (Fig. 1). A subsample was taken by a ran- dom stratified method from each sample for biological examination. In total, 776 individuals were analyzed. For each fish, the TW (0.1 g) and the weight of the gonads (0.01 g) were measured, and sex and stage of matu- ration were ascertained macroscopi- cally. The latter was classified as fol- lows: I = immature; II = resting; III = ripe; IV = ripe and running; V = spent. Sagittal otoliths of the fish were ex- tracted, cleaned, and stored dry. The length-frequency distribution of indi- viduals in catches was calculated. Data were pooled for 1992-95. Lorenzo and Pa|uelo: Biology of Promethichthys prometheus 93 « CANARY ISLANDS J) -29" f-' Lanzarote LaPalmaX / TentT\fe y^-p \ 1 Gran Canaria Fueneventura ^_^ -28" Gomera W ^K 1^ Zt'i,o 05=3.84. Quarter Males Females Sex ratio X^ 3/92 16 23 1.44 1.25 4/92 13 35 2,69 10.08* 1/93 13 37 2,85 11.52* 2/93 24 37 1,54 2.77 3/93 24 34 1,42 1.72 4/93 14 38 2,71 11.07* 1/94 29 51 1,75 6.05* 2/94 36 49 1.36 1.98 3/94 39 50 1,28 1.35 4/94 24 44 1,83 5.88* 1/95 31 59 1,90 8.71* 2/95 19 34 1,79 4.24* tern was recorded for both sexes. Highest values oc- curred between April and September, peaking dur- ing June-July. From October to March the values were low. No significant difference in length at sexual ma- turity was found between males and females latest. immature individuals was from 36 to 38 cm. The parameters of the total length to total weight rela- tionship for males and females separately, and for the population as a whole, are given in Table 4. No significant difference in the allometric coefficient was found between males and females (^-test, t- <=1. 31<0.05) and brown rockfish {F=1.15, P>0.05) there was no significant difference in the amount of variance explained by estimating sex-specific values for the three param- eters. Thus, for each species growth of both sexes is adequately depicted by a single growth curve. The von Bertalanffy parameters for grass rock- fish wereL.= 51.3 cm, ^= 0.11, 1 25 Figure 1 Von Bertalanffy growth curves for grass rockfish (S. rastrelliger) and brown rockfish {Sebastes auriculatus) from southern California. L = total length in centimeters; and a and b - constants. Parameters were estimated with Fishparm v. 3.0 (Saila et al. 1988) by using Marquardt's algorithm for nonlinear least squares parameter estimation. To test for differences in the relationships between sexes, male and female data within a species were modeled separately. Each relationship was then logj^- transformed to create a linear equation, and slopes of the male and female models were compared by using Student's /-test. This process was repeated with gonad weight subtracted from total body weight, to Love and Johnson: Life history of Sebastes rastrelllger and S. auriculatus 103 3,500 -| - grass rockfish ^ , 3,000 - o Male / - X Female / 2,500- W= 0.045 X L2" ^V - r^ = 0.98 x_o/ 2,000 - / - X Xrf / 1,500- P - iS^C 1,000 - - Sjiyo 500- X JM^^ S £ r\ Ol 0- ' i" T^i 1 1 I 1 < 1 1 1 1 c ) 10 20 30 40 50 60 s o F 3,000 - - brown rocWish 2,500 - o Male X y X Female x/ W= 0.044 xL2" x/ 2,000 - r^ = 0.94 X y - o ^ 1.500" oj K^ - 0 ^2gB2 1.000" 500 - 0 - ( 1 1 ■ 1 1 1 1 1 1 3 10 20 30 40 Total length (cm) Figure 2 1 1 1 1 50 60 Length-w« ight relationships for grass rockfish (S. rastrelllger) and brown rockfish (5 ^ebastes auriculatus) from southern California. determine if any difference was simply an artifact caused by the larger female gonads. For grass rockfish, 86 males and 73 females were measured and weighed. Although female grass rock- fish tended to be heavier than males at a given length (T-test, ^=2.21, P<0.05), the difference was not evi- dent when gonad weight was subtracted from total body weight (T-test, ^=1.66, P>0.05), and we there- fore have combined these data into a single figure (Fig. 2). A total of 116 male and 102 female brown rockfish were sampled. For brown rockfish, there was no sig- nificant difference in weight at a given length between males and females (T-test, ^=0, P>0.05); we have also combined these data into a single figure (Fig. 2 ). Length and age at first maturity For these analyses, we examined 53 female and 64 male grass rockfish and 135 female and 129 male brown rockfish. Both male and female grass rock- fish matured over relatively narrow length and age intervals (Table 1; Fig. 3). Both sexes began to ma- ture at 22 cm and all fish were mature by 28 cm. 104 Fishery Bulletin 97(1), 1999 A A A A A Males* *9®50% mai FemalesA 50% mat :3.5yr = 3.7yr Id E c g '■c o a. ~\ I — I — r 2 3 4 5 6 Age (yr) ~1 I I r 7 8 9 10 I T I [ 1 1 1 1 1 1 r 20 21 22 23 24 25 26 27 28 29 30 Total length (cm) Figure 3 Age-maturity and length-maturity and relation- ships for male (n=88l and female in=14) grass rockfish (Sebastes rastrelliger) taken in the southern California Bight, including estimated ages and lengths at which SO'/f of the fish were mature. Solid and dashed lines represent pre- dicted relationship. Fifty percent of males were mature at about 24.5 cm, and fifty percent of females at 24 cm. Male grass rock- fish matured from 2 to 5 yr, females from 3 to 5 yr For brown rockfish, there was also little difference between sexes in either length or age at first matu- rity (Table 1). Males matured between 19 and 29 cm, females between 21 and 32 cm (Fig. 4). For males, 50% maturity occurred at about 25 cm, for females at 26.4 cm. Age at first maturity for both male and female brown rockfish occurred as early as 3 yr, and all fish were mature by 6 yr (Fig. 4). Table t Maximum-likelihood estimates for the parameters of the logistic equation relating proportion mature to lengths and ages of grass and brown rockfishes. Predictive length (/„ j^) and age i.Age^^„) at 50'7t maturity and mean square error (MSE) of the logistic equation are also presented. Species Length a b 'o.50 (=™' r2 Grass rockfish Male Female Brown rockfish Male Female 0.83 0.73 0.38 0.62 20.62 17.49 9.47 16.36 24.5 24.0 25.0 26.4 0.97 0.97 0.85 0.99 Species Age a b ^^0.50 r2 Grass rockfish Male Female Brown rockfish Male Female 1.52 1.95 1.57 1.79 5.36 7.20 6.05 7.51 3.5 3.7 3.9 4.2 0.93 0.99 0.99 0.99 ary (Fig. 5). Females with spawned ovaries were found only during February and March, after which followed a 6-month resting stage. Vitellogenic ova- ries were common from December to February. Ovary indices remained at a minimum throughout spring, summer, and early fall, averaging about 0.3% of body weight, then increased to a peak of 6.8% in Decem- ber ( Fig. 6 ). Testes indices were also low during spring and summer (averaging less than 0.1%), rising to a peak of 0.2% in December (Fig. 6). Brown rockfish spawned from January to August, reaching a peak in January (Fig. 5). Although females with resting stage ovaries were found from April to November, most occurred during the summer. We found vitellogenesis-stage ovaries during most months, particularly between fall and spring. For females, gonosomatic indices were lowest (averag- ing 0.6%) from June to November and peaked dur- ing January at 7.1% (Fig. 6). Testicular indices were lowest from March to August (averaging 0.15'^ ) and peaked in October at 0.62% (Fig. 6). Spawning seasons Grass rockfish released larvae from January to March, reaching a peak in spawning during Janu- Fecundity The relation between grass rockfish fecundity and to- tal length (Fig. 7) was best described by the function Love and Johnson: Life history of Sebastes rastrelliger and 5 auriculatus 105 1 - •J 0.8- t 0.6- 1 0.4- A Males • 7 *9«50%mat = 3-9y^ 0.2- Y Females A r\ . . J *9^50%mat = ^2y^ OT^T'IIIIIIII 2 0123456789 10 1 Age (yr) E c o c o Q. O ^ . ~ ^ ^ 4. 1 - fl 0.8- //A 0.6- A •/ ' 0.4- / ^ Males* / /• ■r'-50%ma, = 25cm 0.2- a y4 / Females A O — - ^y^/A ^'-50%ma. = 26.4cm \J 10 15 20 25 30 35 Total length (cm) Figure 4 Age-maturity and length-maturity relationships for male (n = 120) and female (n = 149) brown rockfish iSebastes auriculatus) taken in the southern Califor- nia Bight, including estimated ages and lengths at which 50'7r of the fish were mature. Solid and dashed lines represent predicted relationship. F = aL\ where F = number of eggs in thousands; L = total length in centimeters; and a and b = constants. The value of the parameters a and b were estimated by fitting the linear function log F = log a + b Log L by least squares. Estimated fecundity ranged from 80,000 eggs for a 26 cm TL individual to about 760,000 for one 46.5 cm long. Discussion al., 1990), although compared to many other groups, their growth rates are quite low (Beverton and Holt, 1960). The largest species of rockfishes tend to grow slowest (y^=0.05, rougheye rockfish, S. aleutianus) and dwarf species exhibit relatively rapid growth (k-OAb, halfbanded rockfish, S. semicinctus). How- ever, between these extremes there is no relation between maximum size and growth rate. Grass rock- fish and brown rockfish growth rates are near the mean for the genus (Love et al., 1990). However, what was usual for the rockfishes was a lack of sexual di- morphism in growth rates. In most species, females grow faster than males, particularly after maturity. Compared with most other eastern Pacific rock- fishes, both species are short-lived. Of the 38 species for which we judge that fairly accurate maximum ages are known, only eight had life spans that were similar to or less than grass and brown rockfishes (Love et al.^). Most species lived to more than 30 years, a majority to more than 40 years. Short-lived species tend to be dwarf taxa, such as shortbelly (S. jordani), Puget Sound (S. emphaeus), squarespot (S. hopkinsi), and honeycomb (S. umbrosus) rockfishes. For brown rockfish, there appears to be little dif- ference in age and length at 50% maturity throughout much its range (Table 2). Although fish off central and northern California are older and larger at 50% matu- rity, those from Washington State and southern Cali- fornia are almost identical. Latitudinal differences in length or age of maturation, or both, occur in a number of rockfish species (e. g. splitnose rockfish, S. diploproa; widow rockfish, S. entomelas; yellowtail rockfish, S. flavidus; black rockfish, S. melanops; bocaccio, S. paucispinis) with fishes from higher latitudes gener- ally maturing when larger or older or both (Wyllie Echeverria, 1987; Love et al., 1990; Field, 1984). How- ever, this difference does not occur in greenstriped rock- fish, S. elongatus (Love et al., 1990). Although it is likely that environmental parameters (e.g. water tempera- ture, food availability, growth seasons) play a role in this phenomenon, it is not yet clear how they interact. With 50% mature at age 4, both grass rockfish and brown rockfish mature at a young age compared with many rockfishes. Off California, most species reach 50% maturity at between 4 and 8 years (Wyllie Echeverria, 1987; Love et al., 1990); thus these two species fall in the younger part of that range. Dwarf species, such as shortbelly rockfish iS. jordani), Puget Sound rockfish (S. emphaeus) and stripetail rockfish (S. saxicola) ma- ture earliest, often when 2 or 3 years old (Moulton, 1975; Wyllie Echeverria, 1987; Love et al., 1990). At the other extreme, the largest and more northerly species, such In the northeast Pacific, rockfish growth rates, as measured by the term k, are quite variable (Love et * Love, M. S., M. Yoklavich, L. Thorsteinson, J. Butler. 1998. guide to the rockfishes of the northeast Pacific. In prep. 106 Fishery Bulletin 97(1), 1999 grass rocMish brown rockfish 100- 8 8 12 6 10 8 - 7 9 6 8 8 14 4 36 23 9 9 9 12 9 11 5 80- 80- 60- Resting 60- 40- 20- 1. < < 40- 20- I.I h Jar Mar May Jul Sep Nov ° 1 1 1 1 1 1 1 1 1 l~l 1 Jan Mar May Jul Sep Nov loo- 100-1 se - 80- 60- Vitellogenesis 60- 40 -J §) 20- S 0- 1 i 40- 20- 1 " "l 1 1 1 1 1 1 1 1 1 1 1 1 .5 Jan Mar May Jul Sep Nov 0-t*T* Jan 1 1 1 1 1 1 1 1 1 1 1 Mar May Jul Sep Nov c 100-1 100-1 o a 80- Q- 80- 60- Eyed 60- 40- 20- 1 NO DATA -\ 40- 1. -■ - - 1 1 1 1 1 1 1 1 1 1 1 Jan Mar May Jul Sep Nov ° II 1 1 1 1 1 1 IT — r Jan Mar May Jul Sep Nov 1 100- 100- 80- 80- 60- 40- 20- Spawned .11 ! 60- 40- 20- 1 "1 1 1 1 1 1 1 1 1 1 1 1 1 Jan Mar May Jul Sep Nov ° 1 1 1 1 1 1 1 1 1 1 1 1 Jan Mar May Jul Sep Nov Month Month Figure 5 Percent composition by month of four gonad condition stages for female grass | rockfish iSebastes rastrelliger) and brown rockfi sh iSebastes aiinculatusttaken in the southern California Bight. The total number of females examined per | month is listed at the top of each month in the "resting" section. Table 2 Age and length at 50'? maturity for male and female brown rockfish by area. Data from Washington ( 1978) and from northern-central California from Wyllie Echeverria ( 1987). Lengths of Washington length to total length based on Echeverria and Lenarz (1984). State from Washington et al. fish are converted from fork Sex Washington Northern- central CA Southern CA Male Female 4 yr. 24 cm 4 yr 26 cm 5yr, 5yr, 31 cm 31 cm 4 yr, 25 cm 4 yr, 27 cm as yelloweye rockfish (S. ruberrimus) and rougheye rockfish iS. aleutianus) may take 20 years or more to mature ( McDermott, 1994; O'Connell'"'). O'Connell. T. 1994. Alaska Department of Fish and Game. 304 Lake St., Room 103, Sitka, AK 99835. Personal commun. Love and Johnson: Life history of Sebastes rastrelliger and 5. auriculotus 107 0.9 -1 0.6- r9 Grass Rockfish O Male X Female * Jan Mar May Jul Sep Nov CO O » 3 B 0.8- 0.6- 0.4- 0.2- rIO Brown Rockfish O Male X Female ^-^:^-.-^-.' -6 -4 -2 I I I I I I I I I I r Jan Mar May Jul Sep Nov Month Figure 6 Seasonal changes in the gonosomatic index (GSI-gonad weight as a percent of total body weight) of male and female grass rockfish iSebastes rastrelliger) and brown rockfish {Sebastes auriculatus). No samples of grass rockfish were taken during November and no samples of male brown rockfish were taken in Febru- ary and September. Vertical lines indicate 95'7t confidence intervals of the mean. Along the northeast Pacific, the vast majority of rockfish spawn in the late winter and early spring and, as with length and age at maturation, many rockfish species exhibit a latitudinal trend in spawn- ing season. Parturition often starts earlier or is more prolonged (or both) in the southern part of a species' geographic range (O'Connell, 1987; Wyllie Echever- ria, 1987; Barss, 1989; Love et al., 1990). The spawn- ing season for brown rockfish from Puget Sound to southern California appears to follow this pattern. Spawning occurs in Puget Sound in June (Washing- ton et al.''), in December-January and May-June in northern-central California (Wyllie Echeverria, 1987) and in January-August in southern California. 108 Fishery Bulletin 97(1), 1999 900- 800- • / _ 700 T grass rockflsh / g' 600- F = 5,267 xL3»<» / § 500- H = 0.95 / • 1 400- i 300- 200- 100- ^^^^ 1 1 1 1 1 1 1 1 1 1 1 1 0 10 20 30 40 50 60 Total length (cm) Figure 7 Fecundity-total length relationship for eight grass rock- fish iSebastes rastreltiger) taken in the southern Califor- nia Bight. Acknowledgments We thank J. Harding, W. Golden, A. Amman, and L. McDonald for help in collecting some of the speci- mens and A. Brooks for providing analytical assis- tance. As always, L. Thorsteinson was very support- ive of our work. This work was conducted through a cooperative agreement with the Biological Resources Division, U. S. Geological Survey, contract number 1445-CA- 0995-0386. Literature cited Adams, P. B. 1992. Brown rockfish. /n W. S. Leet, C. M. Dewees and C. W. Haugen (eds. ), California's living marine resources and their utilization, p. 127. California Sea Grant Extension Publication. UCSGEP-92-12. Barss, W. H. 1989. Maturity and reproductive cycle for 35 species from the family Scorpaenidae found off Oregon. Ore. Dep. Fish Wildl., Inf Rep. 89-7, 36 p. Beverton, R. J. H., and S. J. Holt. 1960. A review of the life span and mortality rates of fish in nature and their relation to growth and other physi- ological characteristics. In G. E. W. Wolstenholme and M. D. Connor (eds.), Ciba foundation colloquia on ageing, vol. 5, The lifespan of animals, p. 142-177. Little, Brown and Co.. Boston, MA. Echeverria, T., and W. H. Lenarz. 1984. Conversions between total, fork and standard lengths in 35 species of Sehastea from California. Fish. Bull. 82:249-251 Eschmeyer, W. N, E. S. Herald, and H. Hammann. 1983. A field guide to Pacific fishes of North America, from the Gulf of Alaska to Baja California. Houghton Mifflin Co.. Boston, 336 p. Feder, H. M., C. H. Turner, and C. Limbaugh. 1974. Observations on fishes associated with kelp beds in southern California. Calif Dep. Fish Game, Fish Bull. 160. 144 p. Field, L. J. 1984. Bathymetnc patterns of distribution and growth in three species of nearshore rockfish from the southeastern Gulf of Alaska. M.S. thesis, Univ. Washington, Seattle, WA, 88 p. Grossman, G. D. 1986. Food resource partitioning in a rocky intertidal fish assemblage. J. Zool., Lond. (B) 1:317-355. Gunderson, D. R. 1977. Population biology of Pacific ocean perch, Sebastes alutus. stocks in the Washington-Queen Charlotte Sound region, and their response to fishery. Fish. Bull. 75:369- 404. Gunderson, D. R., P. Callahan, and B. Goiney. 1980. Maturity and fecundity of four species of Sebastes. Mar Fish. Rev 42(3-41:74-79. Lea, R. N. 1992. Rockfishes: overview. In W. S. Leet, C. M. Dewees and C. W. Haugen (eds.). California's living marine re- sources and their utilization, p. 114-116. California Sea Grant Extension Publication, UCSGEP-92-12. Love, M. S., P. Morris, M. McCrae, and R. Collins. 1990. Life history aspects of 19 rockfish species (Scorpaeni- dae: Sebastes) from the southern California Bight. U.S. Dep. Commer.. NOAA Tech. Rep. NMFS 87, 38 p. Love, M. S., and W. Westphal. 1981. Growth, reproduction, and food habits of olive rock- fish, Sebastes serranotdes, off central California. Fish. Bull. 79:533-545. Matthews, K. R. 1990. A comparative study of habitat use by young-of-the- year, subadult, and adult rockfishes on four habitat types in central Puget Sound. Fish. Bull. 88:223-239. McDermott, S. F. 1994. Reproductive biology of rougheye and shortraker rockfish. Sebastes aleutianus and Sebastes borealis. M.S. thesis. Univ. Washington, Seattle, WA, 76 p. Miller, D. J., and R. N. Lea. 1972. Guide to the coastal marine fishes of California. Calif Dep. Fish and Game, Fish Bull. 157, 235 p. Moulton, L. L. 1975. Life history observations on the Puget Sound rock- fish, Sebastes emphaeus (Starks, 1911). J. Fish. Res. Board Can. 32:1439-1442. O'Connell, V. M. 1987. Reproductive seasons for some Sebastes species in southeastern Alaska. Alaska Dep. Fish and Game, Infor- mational Leaflet No. 263, 21 p. Quast, J. 1968. Observations on the food of the kelp-bed fishes. In W. J. North and C. L. Hubbs (eds.). Utilization of kelp-bed resources in southern California, p. 109-142. Calif Dep. Fish Game. Fish. Bull. 139. 264 p. Saila, S. B., C. W. Recksiek, and M. H. Prager. 1988. Basic fishery science programs. Developments in aquatic and fishery science, vol. 18. Elsevier Scientific Publishing Co., BronxviUe, NY. SVSTAT. 1992. Statistics, version 5.2 edition. SVSTAT Inc.. Evans- ton, IL, 724 p. Love and Johnson: Life history of Sebastes rastrelliger and S aunculatus 109 von Bertalanffy, L. 1938. Aquantitative theory of organic growth. Hum. BioL 10:181-213. Westrheim, S. J. 1975. Reproduction, maturation, and identification of larvae of some Sebastes (Scorpaenidae) species in the northeast Pa- cific Ocean. J. Fish. Res. Board Can. 32:2399-2411. Wyllie Echeverria, T. 1987. Thirty-four species of CaUfornia rockfish: maturity and seasonaHty of reproduction. Fish. BulL 85:229-240. Yoshiyama, R. M., C. Sassaman, and R. N. Lea. 1986. Rocky intertidal fish communities of CaUfornia: tem- poral and spatial variation. Env. Biol. Fish. 17:23-40. 110 Abstract.— Three new species of hag- fish (Myxinidae, Eptatretus) are de- scribed from the Galapagos Islands. Ecuador. These are the first myxinids known from this region and the first species of Eptatretus with five, six. and eight gill pouches reported from the eastern Pacific. A key to their identifi- cation is presented. Three new species of hagfish (Myxinidae, Eptatretus) from the Galapagos Islands Charmion B. McMillan Marine Biology Research Division, 0202 Scripps Institution ol Oceanography La Jolla, California 92093 E-mail address charmcmig)|uno com Manuscript accepted 7 April 1998. Fish. Bull. 97:110-117 (1999). The Galapagos Islands consist of about ten principal islands and over 100 smaller ones on the equator about 500 miles off the coast of Ec- uador. The biodiversity found in these islands by Darwin (1896) is also expressed in the marine fauna (McCosker, 1997); three new species of hagfish were found in only eight specimens from four trap sets (Fig. 1 ). An exploration off South America and the Galapagos Islands in 1891 by the U.S. Fish Commission Steamer Albatross reported only one species of hagfish, Myxine circifrons Garman ( 1899), taken off the Gulf of Panama. Several species of Eptatretus with from nine to four- teen gill pouches have since been reported from the eastern Pacific coast (Wisner and McMillan, 1990), and one species with seven gills, E. laurahubbsae McMillan and Wisner (1984), from the Juan Fernandez Islands, Chile. The new species de- scribed below are the first hagfish reported from the Galapagos Is- lands, and the first Eptatretus with five, six, and eight gill pouches known from the eastern Pacific. Until the formation of the Panama- nian land bridge there was a long- standing connection between the Caribbean and eastern Pacific, which provided a passage for hag- fish to move into the eastern Pacific from the Caribbean. These new spe- cies may be more closely related to the Eptatretus with five to eight gills found in the Caribbean and western Atlantic than to any cur- rently known from the eastern Pa- cific. Future collecting efforts along the coast of Ecuador may reveal Eptatretus similar to those found off the Galapagos Islands; however, until further material is available and genetic studies are made for comparison, we can only speculate on possible origins of these Gala- pagos hagfishes. Although the body color of most species of Eptatretus is brown to black, Fernholm ( 1991 ) described a species of Eptatretus on the basis of one specimen with a highly unusual pink body, stating that its color was probably caused by diet. Of the eight specimens reported here, the seven larger ones are dark purplish- brown to black or dark gray where the slime has not been removed. The smallest specimen (about 142 mm), possibly an albino, is ivory to light tan. Not enough is known about the early development of hag- fish to determine if this tiny speci- men would have become darker with age. Albinism has been re- ported by Dean ( 1903) and Jensen ( 1959), and I have collected one, an adult E. deani (Evermann and Goldsborough, 1907), which was pinkish-white when alive and light tan color in formalin. Head grooves are present near the eyespots of the seven large specimens, but not on the smallest hagfish. These grooves, found in many other species of Eptatretus (McMillan and Wisner, 1984) and once thought similar to lateral lines, McMillan: Three species of hagfish from the Galapagos Islands 111 are now believed to have no sen- sory purpose. Because they are not considered a species-specific character and have no taxo- nomic value for the present study, they are not included in the species descriptions. Counts of cusps and total slime pores are somewhat simi- lar for these new Eptatretus spe- cies, but there are significant differences in the multicusps, numbers of gill pouches, and prebranchial slime pores which readily distinguish each of them. Some differences are found in the white patterns on the face and barbels of the seven pig- mented specimens (Fig. 2), but color cannot be assigned major importance as a species charac- ter with such a small sample size. Material and methods All new specimens for this study were collected by John E. Mc- Cosker et al using the research submersible Johnson Sea-link (JSL), on the California Academy of Sciences and Harbor Branch Oceanographic Institute Galapagos Expedition during November 1995. The three new species collected by the California Academy of Sci- ences and described below were among eight speci- mens taken in four trap sets; additional specimens (nontypes) of Galapagos Eptatretus were given to the Institute Nacional de Pesca, Guayaquil, Ecuador, and were not available for study. Collection data and dis- position of specimens are listed in the treatment of each new species. Institutions in which type speci- mens have been deposited are the California Acad- emy of Sciences, San Francisco (CAS), United States National Museum, Washington, D.C. (USNM), and Scripps Institution of Oceanography, La Jolla, Cali- fornia (SIO). McCosker^ described the method of capture as fol- lows: "Hagfish were collected with a small, galva- nized wire commercial two-fyked minnow trap, about 30 by 50 cm, with 2 cm openings at each end. The Figure 1 Principal islands of the Galapagos Archipelago, showing collection locations of the three new species of Eptatretus: ■ = £. grouseri; ♦ = £ mccoskeri; and A = E. wisneri. ^ McCosker, J. E. 1996. California Academy of Sciences, ~=r Golden Gate Park, San Francisco, CA, 941 18-4599. Personal commun. traps were baited with fish or molluscs, or both, weighted and set with a pinger and short float line to be retrieved on the next dive [about 8 h later]; all traps were set among boulders, often on steep slopes." McCosker further reported that every set resulted in hagfish so elusive that many escaped traps dur- ing or after retrieval. Hagfish were also seen, but not taken, at 267 m off East Seymour Island, at 274 m off Isla Floreana (Charles Island), and off Isla San Salvador (James Island) at 884 m over sand bottom, at 6.54°C. Although no study material resulted from these sets, the reports furnish information on the little-known habitat of these benthic fish and indi- cate locations for further collecting effort. Methods of measuring and counting follow those of McMillan and Wisner (1984) and Wisner and McMillan (1995). Counts of gill pouches (GP), gill apertures (GA), and cusps are noted for both sides, but body proportions and slime pore counts are for the left side only. Low power magnification was re- quired to discern head grooves and fusion of multicusps. Most of the specimens are twisted and previously cut, making some measurements and counts difficult; all specimens were measured several times and averages 112 Fishery Bulletin 97(1), 1999 B D E G i Figure 2 Frontal aspect showing pigmentation on face and barbels of three new species of Eptatretus from the Galapagos Islands (Al holotype of £. grousen (CAS 864281; (Bi holotype of £. wisneh (CAS 86429); (Cl paratype of E. wisnen (SI097-76) (D) holotype of£. mccoskeri fCAS 864311; lE) and (F) paratypes of£. mccosken (SIO 97-75; and (G) paratype of £. mccoskeri (USNM 344905). used to calculate body proportions in thousandths of total length (TL). About 6-9 mm of shrinkage was noted in the total lengths of the seven larger specimens be- tween 24 March 1996 and 20 March 1997. Some shrink- age occurs in all specimens on preservation, mainly in the trunk length, and apparently slightly more when ethanol is the preservative used, as in this case. Se- lected body proportions and counts of slime pores and cusps are shown in Table 1. Because variations in color patterns are often noted in large collections, color is not usually considered of much importance. However, the two species with eight gills have distinct differences in the white pat- terns on their faces and barbels that may provide a useful character in distinguishing between them. The "face" is the anteriormost part of the specimen, the ventral aspect of the head, delineated by three pairs of long, slender protruberances thought to be sen- sory organs and commonly referred to as barbels. The first two pairs are above and below and adjacent to the nasopharyngeal duct, which is the intake open- ing for water to the pharynx. The third pair lies just outside another set of short, fleshy protruberances on each side of the mouth. McMillan: Three species of hagflsh from the Galapagos Islands 113 Table 1 Selected measurements and counts of slime pores and cusps of three new species of Eptatretus from the Galapagos Islands. Species name followed by number of specimens in parentheses; total length of holotype followed by TL of paratype(s) in parenthe- | ses; body proportions in thousandths of TL, average followed by values for all specimens. Measurements E. grouseri (2) E. mccoskeri (4) E. wisneri (2) Total length in mm 380 (142) 320 (283,298,300) 360 (328) Prebranchial length 225(211-239) 249(237-262) 213(194-229) Branchial length 71 (63-79) 97(93-101) 105(100-111) Trunk length .544 (535-553) 494 (487-500) 516(503-28) Tail length 160(158-162) 162(156-177) 168(167-171) Body depth withVTF 58(49-68) 100(94-106) 88(85,92) Body depth without VFF 54(42-66) 100(94-106) 87 (85-89) Body depth at cloaca 54(48-61) 83 (78-88) 72(72-73) Tail depth 63(63-63) 93 (87-102) 77(76-78) Slime pore counts (holotypye followed by para type! s ) in parentheses): Prebranchial pores 12(13) 15(14,14,14) 9(9) Branchial pores 4(5) 7(7,7,7) 7(7) Trunk pores 46(44) 42 (40,40,41) 46 (47) Tail pores 15(14) 10(10,11,12) 14(13) Total pores 77(76) 74(72,72,74) 76(76) Cusp counts, left, right (holotype followed by paratype(s) in parentheses: Multicusps. ant./post. 3,3/2,2 3,3/3,3 3,3/2,2 Unicuspslanterior rows) 8,9(9,9) 10,10*0,9-9,10) 9,9 (9,9) Unicusps(posterior rows) 8,8(8,8) 9,10(9,9-9,10) 8,8(8,8) Total cusps 44(44) 51 (48,50,51) 44(44) called "labial barbels," which are immediately adja- cent to the oral opening, very similar in size and shape in all hagfish, and are not considered species signifcant. The other three pairs of long, slender bar- bels are usually measured, and sometimes used as a species character. Because many of the barbels on these specimens were curled and accurate measure- ments difficult to repeat, this character has not been included in Table 1. Multicusp pattern, and counts of GP and slime pores have proved to be more accu- rate and less variable intraspecifically than color or barbel size, and thus more important as species char- acters. The most significant features of taxonomic im- portance in this study are the counts of gill pouches, prebranchial slime pores, and multicusps. "First" and "last" refer to anteriormost and posteriormost, and tail pore counts are the sum of cloacal and caudal slime pores; other terms used are from Wisner and McMillan ( 1990, 1995). The water passes fi-om the the nasophary- geal duct through the pharynx, through each afferent branchial duct (ABD) to the gill pouch, and exits through each efferent branchial duct (EBD) to its cor- responding external opening (GA). Any excess water passes out of the body through the slightly larger duct known as the pharyngocutaneous duct (PCD), which is ordinarily confluent with, or occasionally just poste- rior to the last GP on the left side. The numbers of gill apertures are not shown in the table because the num- ber of GA is normally the same as that of GP in the genus Eptatretus; however, GA count is given in each species description and in the key because the aper- tures are readily counted by external examination and thus provide a useful character for identification in the field. In most Eptatretus, including the three species described in the present study, the number of bran- chial pores is one less than the number of gill apertures, and each pore is slightly below and poste- rior to each GA anterior to the PCD. The next slime pore, usually found above and behind the PCD, is the first in the trunk pore series. Key to the species of Eptatretus of the Galapagos Islands la Five or six gill pouches and apertures each side; multicusps 10 (3 in each anterior row and 2 in each posterior row) total cusps 44; prebranchial slime pores 12-13, trunk pores 44-46, tail pores 14-15, total pores 76-77 E. grouseri, new species 114 Fishery Bulletin 97(1), 1999 lb Eight gill pouches and apertures each side 2 2a Prebranchial slime pores 14-15, trunk pores 40-42, tail pores 10-12; total pores 72-74, total cusps 48-51, multicusps 12 (3 in each row); nasal barbels completely white; face mostly dark with a small white area around mouth and on labial barbels E. mccoskeri, new species 2b Prebranchial slime pores 9; trunk pores 46-47, tail pores 13-14; total pores 76, total cusps 44, multicusps 10 (3 in anterior, 2 in posterior rows), nasal barbels dark with occasional white tips; face, mouth, and base of labial barbels nearly all white E. wisneri, new species Eptatretus grouseri new species Holotype CAS 86428, female, 378 mm TL, taken at 00°14.6'S, 91°26.6'W, in a minnow trap at 2370 ft [722 m], 17 November 1995. Paratype SI097-77 (formerly CAS 86428), 142 mm TL, (juvenile, sex not determined), taken with the holotype. Diagnosis Five or six gill pouches and apertures each side, last GA confluent with PCD on left side. Multicusps 3 in anterior and 2 in posterior rows, 9 unicusps in each anterior and 8 in each posterior row, total cusps 44; slime pore counts: prebranchial 12- 13, branchial 4-5, trunk 44-46, tail 14—15, total slime pores 76-77; ventral fmfold (VFF) vestigial, caudal fmfold (CFF) well developed. Description Both specimens have a bluntly rounded rostrum; barbels well developed, first two pairs about equal, third pair slightly longer; greatest body depth in trunk region about the same as tail depth, both only about 6 to 7 percent of TL; prebranchial length about 23 percent of TL, trunk length about 54 per- cent of TL. Although the two specimens differ greatly in size and color, counts of pores and cusps are very similar (Table 1). The holotype has tiny round eggs less than one mm in diameter, no ellipsoidal devel- oping eggs and no tissue indicating previous large eggs. Body color brownish-black, head region dark brown; eye spots prominent, nearly round; face dark with small white area around mouth and on tips of barbels (Fig. 2A); VFF vestigial, with pale margin on posterior third; cloacal opening white; GA widely spaced with branchial slime pores posterior to and just slightly below each aperture; all GA on the right side and all but the sec- ond on the left side have white margins. The second gill pouch on the right side of the holotype is either undeveloped or degenerated; it appears only as thick- ened muscle tissue connecting the ABD from the phar- ynx with the EBD leading to the external GA. Only one GP lies along the tip of the dental muscle; the bran- chial aorta splits after the fifth (last) GP. The paratype has six normally rounded GP on each side, the first two along the dental muscle with the aorta branching after the fifth GP. Possibly an al- bino, this tiny specimen is an unusual ivory to light tan body color, the prebranchial area slightly lighter with eye spots barely discernible. Also, the area around the mouth and cloaca are lighter than the body; the trunk region is just dark enough to show a white margin on the VFF. The slime pore openings are very small, but easily detected because the sacs bulge just under the thin skin. The specimen is so young that it was not possible to determine gender even with magnification. The total cusp count only differs from that of the holo- type by one, but they are very small and the multicusp fusion pattern is too indistinct to discern the number of fused cusps in the posterior rows with any certainty. Etymology This species is dedicated to my son David "Grouser" McMillan, a Chief Engineer in the U.S. Merchant Marine, for his continued encourage- ment of my hagfish studies and for his knowledge and love of ships and the sea. Distribution Known only from Galapagos Islands, Ecuador; both specimens were collected in JSL dive 3958 at 2370 ft [722 m] in Bolivar Channel, off'Cabo Douglas, in a minnow trap set on a steep, sediment- laden slope of an invertebrate-rich pinnacle of pil- low lava. This was the deepest set in which hagfish were collected on this expedition (Fig. 1). Comments The body proportions differ by about two percent in prebranchial length and depth of body in trunk region, possibly because one specimen is a ju- venile. Pore and cusp counts, which remain the same throughout the individual's life span, are very simi- lar in both specimens. In this tiny paratype the first GP is about 3.5 mm in diameter, with each subse- quent GP decreasing in size posteriorly until the sixth is only about 2 mm across. This difference has also been noted in size of unicusps, in which the last one is frequently much smaller at the end of each row, apparently a normal growth pattern. I have not made any study of changes in body proportions of hagfishes from juvenile to adult stage because very small speci- mens are seldom collected. The capture of this tiny McMillan: Three species of hagfish from tfie Galapagos Islands 115 specimen was possible because of the use of a min- now trap with very small openings. Because of the larger openings of the mjrxinid traps most often used, hagiishes of less than 200 mm length are seldom caught because they escape before a trap reaches the surface. The larger specimen with normal pigmentation has been designated as the holotype (in spite of its one abnormal GP), because its adult size allowed deter- mination of gender and multicusp pattern, as well as more accurate measurements and slime pore counts. These two specimens are assigned to the same species because there is such close agreement in cusp and slime pore counts and most body proportion and because they were collected in the same JSL dive. A difference of one gill pouch is common in large col- lections of Eptatretus having nine or more GP, and this difference may be found occasionally in species having only five or six gill pouches. More specimens will be required to determine the usual number of GP, and whether there are enough other differences to warrant separating into two different species. Discussion Three species of five- or six-gilled Eptatretus from the Caribbean Sea are E. minor and E. multidens Fernholm and Hubbs (1981), and E. mendozai Hensley ( 1985). Although these three spe- cies may have once shared a common ancestor with E. grouseri, they are readily distinguished by their unusual 3/3 multicusp pattern (three fused cusps on both the anterior and posterior rows of cusps). The 3/2 multicusp pattern of E. grouseri commonly oc- curs in most species of Eptatretus: two other five- gilled species with this pattern are E. profundis (Barnard, 1923) from South Africa and E. eos Fernholm (1991) from the Tasman Sea, the latter a distinctive pink color and possessing a tube-shaped, elongated "snout." Both of these species have about 26 prebranchial slime pores, an unusually high num- ber for Eptatretus and twice as many as found in E. grouseri. Four species of Eptatretus with six gill pouches and the 3/2 multicusp pattern are E. burgeri (Girard, 1854) from Japan and Korea, E. hexatrema Barnard (1923) from South Africa, E. longipinnis Strahan (1975) from South Australia, and E. chinensis Kuo and Mok (1994) from the South China Sea. Eptatretus burgeri has a distinctive white mid- dorsal line as well as notably higher prebranchial and total slime pore counts, whereas the other three species have prebranchial pore counts two to three times that of E. grouseri. Although the number of GP is the same, these six species are not considered closely related to E. grouseri because of the signifi- cant differences in prebranchial and total slime pore counts, as well as wide geographic separation. Eptatretus mccoskeri new species Holotype CAS 86431, male, 320 mm TL, taken at 01°06.3'S, 89°06.9'W, in a minnow trap at 704 ft [215 m], 16 Nov 1995. Paratypes All males, taken with the holotype; SI097-75, 2(283,300 mm TL); and USNM 344905, 298 mm TL. Diagnosis Eight gill pouches and apertures each side, last GA confluent with the PCD on the left side; 12 multicusps (3 fused cusps in each row), unicusps 36— 39 (9,9 to 10,10 unicusps in each anterior and 9,9 to 9,10 in each posterior row), total cusps 48-51; total slime pores 72-74, prebranchial 14-15, branchial 7, trunk 40-42, tail 10-12. Ventral fmfold absent or vestigial; caudal fmfold with white margin; head slightly lighter than body color, face dark except for small white area around mouth; barbels all white (Fig. 2, D-G). Description Body color brownish black with head region slightly lighter than body color, face dark ex- cept for small white area around the mouth; eye spots visible, but not prominent; face dark, in sharp con- trast to completely white barbels; prominent white margin around CFF; rostrum bluntly rounded, nearly straight across; nasal orifice large, face wide with the distance between the pairs of barbels about equal to that of their length. First five (4-6) GP lie along the dental muscle; branchial aorta branches at the sixth or seventh GP; first GA high, series curving downward to the PCD near the ventral midline; se- ries of prebranchial slime pores starts far forward; branchial pores are tiny and close to ventroposterior margins of each GA, trunk pore series starts above and behind the PCD, a space about 5-6 mm between last trunk pore and origin of cloaca. Prebranchial length is about one fourth of TL, trunk length about half of TL; greatest body depth in the trunk region about equal to the tail depth, both about 10 percent of TL (Table 1). Etymology This species is dedicated to my friend John E. McCosker, Senior Scientist, California Acad- emy of Sciences, San Francisco, California, for pro- viding this new study material, as well as for his important contributions to marine biology. Distribution Known only from the Galapagos Is- lands, Ecuador; all four specimens were taken in one dive (JSL 3936) from a minnow trap set on sand bot- tom at the top of a seamount southeast of Isla San Cristobal (Chatham Island) at about 215 m, the shal- 116 Fishery Bulletin 97(1), 1999 lowest set where hagfish were found on this expedi- tion (Fig. 1). Comments Because male hagfish rarely outnum- ber females in any collection, it is unusual to find that these four specimens are all males; testes are developed, but no study was done to determine pres- ence of sperm. Although both E. mccoskeri and E. wisneri (described below) have eight GP, the pattern of three fused cusps in both anterior and posterior rows, notably higher numbers of prebranchial pores and total cusps, and the sharp contrast of white bar- bels against the dark face ofE. mccoskeri readily dis- tinguish these two species. Also, the prominent white margin around the entire caudal finfold in E. mccoskeri is much less apparent in E. wisneri, show- ing only as a thin white edge on the tip of the CFF. The numbers of pouches along the dental muscle and position of branching of the branchial aorta also dif- fer slightly in these two 8-gilled species. Three other species of Eptafretus with 8 GP and the 3/2 multicusp pattern are listed below in the discussiom of £. wisneri. Eptatretus wisneri new species Holotype CAS 86429, female, 356 mm.TL, taken at 00=28. 0'S, 91°37.2'W, from a minnow trap at 1848 ft [563 m], 14 Nov 1995, [JSL dive 3952]. Paratype SI097-76 [formerly CAS 86430] , male, 328 mm TL, taken at 00=17. 5'S, 91°38.9'W, from a min- now trap at 1680 ft [512 m], 16 Nov, 1995 [JSL dive 3957]. Diagnosis Eight gills and apertures on each side, last GA on left side confluent with PCD; 10 multi- cusps, pattern of 3 fused cusps in each anterior row and 2 in each posterior row; unicusps 9 each ante- rior row and 8 each posterior row; total cusps 44; total slime pores 76, prebranchial pores 9, branchial pores 7, trunk pores 46-47, tail pores 13-14; face mostly white, including mouth and base of labial barbels; other barbels dark, with occasional white spots or tips (Fig. 2, B and C); eye spots distinct, large and irregularly shaped; VFF vestigial or absent, CFF well developed. Description Body color brownish-black, head slightly lighter, with large, distinct eye spots; gill apertures with white margins; face nearly all white, including area around mouth and base of labial bar- bels; the VYF appears only as a pale line on the pos- terior half of the trunk region and the tip of the CFF has a thin pale margin. The first 3 gill pouches lie along the dental muscle; the branchial aorta branches after fifth GP; the prebranchial length is about 21 percent of TL, trunk length about half of TL, the greatest body depth in trunk region about equal to, or slightly more than the tail depth, both about 8 to 9 percent of TL (Table 1). Etymology This species is named after Robert L. Wisner, my friend and associate at Scripps Institu- tion of Oceanography, for his invaluable assistance with my myxinid research as well as his other con- tributions to ichthyology. Distribution Known only from Galapagos Islands, Ecuador, these two specimens were taken by min- now trap on two different JSL dives, off Cabo Dou- glas, Isla Fernandina (Narborough Island), at 512 and 563 meters (Fig. 1). Comments The holotye is a young female with round eggs of less than one mm, with no ellipsoidal oocytes and without evidence of tissue indicating previous vitellogenesis. The paratype is a male with well-developed testes, but no examination was done for the presence of sperm. One specimen is more brown than black, and the white face patterns are slightly different; also, the barbels of one specimen are partly white and more robust than those of the other ( Fig. 2, B and C ). These slight differences could be due to gender, although pigmentation is highly variable and sexual dimorphism has not been re- ported in hagfish. Barbels are seldom used as a spe- cies character because they are often curled, making accurate measurements difficult, and not reliable as a body proportion because their lengths are so small compared with TL. These two specimens have the same pattern of multicusps and unicusp counts, and slime pore counts are the same or only differ by one. Therefore, until further collecting provides more specimens for comparison, the slight differences in pigmentation patterns and size of barbels are con- sidered insignificant, and the two specimens are de- scribed as one species. Discussion Eptatretus wisneri is readily distin- guished from E. mccoskeri because of its multicusp pattern of 3/2 (three on each anterior and two on each posterior row), lower numbers of total cusps, nota- bly lower number of prebranchial and total slime pores, as well as the darker barbels and the white areas over most of the face. Three other eight-gilled Eptatretus with the 3/2 multicusp pattern are E. octatrema Barnard (1923) from South Africa, E. indrambarayai Wongratana (1983) from the Anda- man Sea (N.E. Indian Ocean), and E. okinoseanus (Dean, 1904) from Japan). However, they are not con- McMillan: Three species of hagflsh from tfie Galapagos Islands 117 sidered closely related to E. wisneri because of dif- ferences in counts, the wide geographical separation, and the fact that this multicusp pattern is common to most species of Eptatretus. Acknowledgments My deepest appreciation is given for the efforts of J. E. McCosker and all persons involved with the California Academy of Sciences-Harbor Branch Oceanographic Institute Galapagos Expedition for making this new study material available. Special thanks are given to R. L. Wisner for valuable help with references and figures, and to R. H. Rosenblatt, H. J. Walker and F. H. Martini for critical review of this manuscript. Thanks are also given to N. D. Holland for providing laboratory space, and to R. M. McMillan for his technical assistance with word processing. Literature cited Barnard, K. H. 1923. Diagnosis of new species of marine Fishes from South African waters. Ann. S. Afr. Mus.l3(pt 8, no.l4):439-445. Darwin, C. 1896. Journal of researches into the natural history and geology of the countries visited during the voyage of H. M.S. Beagle, 1831-1836. 2nd ed. Appleton. New York, NY. Dean, B. 1903. Albinism, partial albinism and polychromism in hagfishes. Am. Natural. 37(437):295-298. 1904. Notes on Japanese myxinoids: a new genus, Para- myxme, and a new species, Homea okinoseana, reference also to their eggs. Jpn. Coll. Sci., Imper. Univ., Tokyo 19(21:1-23. Evermann, B. W,. and E. L. Goldsborough. 1907. The fishes of Alaska. Bull. U.S. Bur. Fish. 26(for 1906):219-360. Fernholm, B. 1991. Eptatretus eos: a new species of hagfish (Myxinidae) from the Tasman Sea. Jpn. J. Ichthyol. 38(2):115-118. Fernholm, B., and C. L. Hubbs. 1981. Western Atlantic hagfishes of the genus Eptatretus (Myxinidae) with description of two new species. Fish. Bull. 79(l):69-83. Garman, S. 1899. The fishes: reports of an exploration off the west coasts of Mexico, Central and South America, and off the Galapagos Islands in charge of Alexander Agassiz by the U.S. Fish Comm. Steamer A/6a?ross during 1891, Lt. Com- mander Z.L. Tanner, U.S.N, commanding. Harvard Mem. Mus. Comp. Zool. 24:1-431. Girard, C. 1854. Family of Myxinoidea. /n Contributions to the fauna of Chile. Proc. Acad. Nat. Sci. Philad., p. 47^9. Hensley, D. A. 1985. Eptatretus mendozai, a new species of hagfish (Myx- inidae) from off the southwest coast of Puerto Rico. Copeia 1985(4):865-869. Jensen, D. 1959. Albinism in the California hagfish. Eptatretus stoutii. Science (Washington, D.C.) 130(3378):798. Kuo, C-H., and H-K Mok. 1994. Eptatretus chtnensis: a new species of hagfish (Myx- inidae; Myxiniformes ) from the South China Sea. Natl. Sun Yat-Sen Univ. Studies 33(4):246-250. McCosker, J. E. 1997. Letter from the field. Galapagos Islands: a half mile down. Pacific Discovery 50(l):42-45. McMillan, C. B., and R. L. Wisner. 1984. Three new species of seven-gilled hagfishes (Myxinidae. Eptatretus) from the Pacific Ocean. Proc. Caiif Acad. Sci. 43(16)249-267. Strahan, R. 1975. Eptatretus longipinnis, n.sp., a new hagfish (family Eptatretidae) from South Australia, with a key to the 5-7 gilled Eptatretidae. Austral. Zool. 18(3):137-148. Wisner. R. L,. and C. B. McMillan. 1990. Three new species of hagfishes. genus Eptatretus (Cyclostomata, Myxinidae) from the Pacific coast of North America, with new data on E. deani and E. stoutii. Fish. Bull. 88:787-04. 1995. Review of new world hagfishes of the genus Myxine (Agnatha, Myxinidae) with descriptions of nine new species. Fish. Bull. 93:530-550. Wongratana, T. 1983. Eptatretus indrambaryai, a new species of hagfish ( Myxinidae ) from the Andaman Sea. Nat. Hist. Bull. Siam Soc. 31(2):139-150. 118 Abstract.— The aim of this project was to investigate the use of strontium as a chemical tag in the dorsal spines of the marine teleost Pagrus aiiratus that would allow the mass tagging of juve- nile fish. Previous studies in which the incorporation of strontium has been ex- perimentally manipulated for the pur- poses of marking have generally con- centrated on freshwater and anadro- mous species. This is the first study to investigate the tagging of spines with strontium, the removal of which is non- destructive. Inductively coupled plasma- mass spectrometry (ICP-MS) was used to measure isotopic concentrations. The dorsal spines of juvenile P. auratus that had been immersed in salt water containing 0.125 g/L SrCU-6H.,0 (5x ambient strontium) and 0.250 g/L (lOx ambient) for five days incorporated *^Sr at levels greater than those in control fish. The strontium signal was persis- tent in spines for at least 36 days and showed no sign of decay during the ex- periment. No effects of the treatments on fish health or growth were detected. Short-term immersion experiments (6 hours to 5 days) indicated that treat- ments of lOx ambient or greater for 4- 5 days were required to tag fish reli- ably with strontium. Natural levels of strontium in the spines of juveniles varied among locations separated by tens of kilometres along the coast of New South Wales. Natural variations in strontium concentrations were not great enough, however, to obscure the differences between tagged and wild fish. It was concluded that strontium immersion is a useful and relatively environmentally safe method of tagging large numbers of small fish. Chemical marking of juvenile snapper, Pagrus auratus (Sparidae), by incorporation of strontium into dorsal spines Morgan J. Pollard Michael J. Kingsford School of Biological Sciences A08 University of Sydney New South Wales 2006, Australia E-mail address (for M J, Kingsford, contact author) MikekiSbio usyd edu au Stephen C. Battaglene ICLARM, Coastal Aquaculture Centre PO Box 438 Honiara, Solomon Islands Manuscript accepted 22 April 1998. Fish. Bull. 97:118-131 (1999). Calcium is incorporated into the calcium carbonate matrix of otoliths and into the calcium phosphate matrix of the skeleton, spines, and scales as fish grow (Francillon- Vieillot et al., 1990). In addition to calcium, trace elements, such as strontium, are also incorporated into the calcified components offish. The chemical similarities of Ca"* and St'^* ions (i,e. similar ionic ra- dius and identical valence) allow strontium ions to act as replace- ments for calcium during the pro- cess of calcification, Ca and Sr con- centrations in calcified components have been explored for their rela- tionships with variations in envi- ronmental conditions (Edmonds et al„ 1989; Gallahar and Kingsford, 1992), such as temperature (Kalish, 1989;Radtkeetal., 1990;Townsend et al,, 1992, 1995) and salinity (Kalish, 1990; Coutant and Chen, 1993; Secor et al., 1995), The manipulation of ambient lev- els of chemicals such as strontium can allow fish to be marked or tagged invisibly. The marking of skeletal structures with isotopes offers a potential method for rapid tagging of large numbers of small juvenile fish. Nonradioactive stron- tium has been investigated, and artificially induced strontium marks have been detected in otoliths (Brown and Harris, 1995; Mugiya and Satoh, 1995; Mugiya and Tan- aka, 1995; Schroder et al,, 1995, 1996; Gallahar and Kingsford, 1996), scales (Ophel and Judd, 1968; Behrens-Yamada and Mul- ligan, 1982; Snyder et al„ 1992), vertebrae (Behrens-Yamada et al,, 1979; Behrens-Yamada and Mul- ligan, 1982, 1987; Schroder et al„ 1995), and opercular bones (Guillou and de la Noue, 1987; Schroder et al,, 1995), The majority of this work, however, has been done on freshwa- ter or anadromous species. The species investigated in this study was the snapper or red sea bream, Pagrus auratus (Pisces; Sparidae), The northern hemi- sphere and southern hemisphere forms (previously Pagrus major and Chrysophrys auratus, respectively) are now considered to be morpho- metrically identical (Paulin, 1990), This highly valued commercial spe- cies supports a very well established intensive aquaculture industry in Japan (Foscarini, 1988; Davy, 1990, 1991; Fukusho, 1991) and shows good potential for aquaculture in Pollard et al : Chemical marking of Pagrus auratus 119 Australia and New Zealand (Battaglene and Bell, 1991; O'Sullivan, 1992; Treadwell et al.M. Commer- cial catch rates of snapper have been on the decline in many areas (Paul, 1982; Gilbert, 1986), and one means of increasing natural populations is by reseed- ing the sea with hatchery-reared larvae or juveniles. Pagrus auratus is one of the targets for this approach in both the northern (Smith and Hataya, 1982; Ishibasi, 1986; Matsuda, 1992) and southern (Smith and Francis") hemispheres. Tagging techniques are required to measure the success of such stock en- hancement measures. Very large numbers of small ju- veniles must be tagged in these studies, and isotope tagging may be one of the more suitable techniques. Tagging techniques have been extensively used for research on Pagrus auratus. Studies involving tag- ging have been used to investigate migration (Crossland, 1976; Jones, 1981; Sakamoto, 1984; Kato et al., 1991), the identification of separate stocks (Sanders, 1974; Moran, 1987), growth rates (Sand- ers and Powell, 1979; Francis and Winstanley, 1989; Tsukamoto et al., 1989; Kato et al., 1991), fishery exploitation and resilience (Crossland, 1980), and the validation of aging with oxytetracycline (Ferrell et al., 1992; Francis et al., 1992). With the exception of fluorochrome markers, such as oxj^etracycline, many conventional tagging techniques are, however, un- suitable for tagging small juveniles or are too en- ergy intensive or expensive for the tagging of large numbers offish. The objective of our study was to determine whether juvenile snapper (Pagrus auratus) could be reliably chemically tagged by the incorporation of strontium into the dorsal spines from immersion in strontium chloride. The specific aims were as follows: 1) to investigate the persistence of chemical tags in the dorsal spines of juvenile snapper following im- mersion of the fish in strontium chloride; 2) to de- termine minimum strontium concentrations and immersion periods required for effective strontium tagging; 3) to determine the feasibility of batch tag- ging with different levels of strontium; and 4) to com- pare the levels of strontium in tagged snapper with naturally occurring levels in wild-caught snapper from different estuaries. These multiple controls were required to ensure that natural variation did not con- found the identification of treated fish. Methods Experimental work was carried out at the New South Wales (NSW) Fisheries Brackish Water Fish Culture Research Station (BWFCRS), Port Stephens. The experimental fish were hatched on 13 November 1993 from wild parent snapper caught off Broughton Is- land, NSW. Rearing procedures and larval charac- teristics may be found in Fukuhara (1985), Lopez ( 1986), Pankhurst et al. ( 1991 ). Battaglene and Tal- bot (1992), and Kingsford and Atkinson (1994). Ex- perimental fish were kept in 2000-litre tanks on a constant flowthrough system of biofiltered oceanic water for the first three weeks, followed by biofiltered estuarine water from Salamander Bay, Port Stephens. From 14 January 1994, 770 juvenile snapper were reserved for the experiments. The strontium salt used was strontium chloride (SrCl,/6H,,0), a relatively cheap, nontoxic and non- radioactive chemical. Strontium is found naturally in seawater at an elemental concentration of approxi- mately 8.1 mg/L (Home, 1969). For simplicity the amounts of strontium chloride added to the saltwa- ter experimental tanks are expressed as multiples of this ambient strontium concentration, as follows: 1) Sr2* five times ambient = 0.125 g/L of SrCl2-6H20; 2) Sr^-^ ten times ambient = 0.250 g/L of SrC^-eHgO; and 3) Sr'-* forty times ambient = 1.000"g/L of SrCl,,-6H20. Preliminary experiments (Pollard-^) showed increased levels of dorsal spine strontium in juvenile snapper that had been immersed in stron- tium chloride solution. Experiment 1 : persistence of strontium The aim of this experiment was to determine whether the strontium marks induced in the dorsal spines of juvenile snapper were persistent over at least a 36- day period. The experiment was initiated on 22 March 1994 when the fish were 129 days old and 68 mm mean fork length (SD ±6.7 mm). The experiment consisted of Sr^* five times ambient, Sr^* ten times ambient, and a control, each with four replicate 100- litre tanks containing eight fish each. Two fish were sampled from each tank at zero, 12, 24, and 36 days following the termination of the five-day exposure period. For comparisons between sample times, treat- ments, and tanks we used a partially hierarchical three-way analysis of variance. 1 Treadwell, R., L. McKelvie, and G. B. Maguire. 1992. Poten- tial for Australian Aquaculture. Research Report 92.2 of the Aust. Bureau of Agricultural and Resource Economics, Canberra, Australia, 81 p. ^ Smith, P. J., and M. P. Francis. 1991. Snapper reseeding in the Hauraki Gulf: scientific considerations. N.Z. MAF Fish- eries Internal Report 172, Wellington, New Zealand, 22 p. ^ Pollard, M. J. 1994. Chemical marking of juvenile snapper {Pagrus auratus: Sparidae) by the incorporation of iron and strontium in the spines, scales and otoliths. B.Sc. Hons. the- sis, School of Biological Sciences, Univ. Sydney, Sydney, Aus- tralia, 119 p. 120 Fishery Bulletin 97(1), 1999 It was possible that Sr could affect fish growth. Thus, the change in mean fish weight per tank between the initial stocking of the tanks and the sampling of the fish remaining after 36 days was compared between treatments by using one-way analysis of variance. Experiment 2: concentration and exposure period The aim of this experiment was to investigate a range of strontium treatments and exposure periods in or- der to determine minimum concentrations and times required for marking the dorsal spines of juvenile snapper. It was also possible that different batches could be identified by different levels of strontium incorporation. The experiment was initiated on 3 March 1994 when the fish were 110 days old and 64 mm mean fork length (SD ±4.1 mm). The 16 treatments com- prised three strontium concentrations (Sr-* five times, ten times, and forty times ambient) matched with each of five exposure periods (6, 12, 24, 36, and 48 hours) and a control treatment. Replication com- prised two 100 litre tanks per treatment, each con- taining six fish. Snapper were sampled seven days after the termination of the chemical treatment. Sta- tistical analysis involved comparisons between treat- ment concentrations, exposure periods, and tanks us- ing an asymmetrical design analysis of variance and the Student-Newman-Keuls (SNK) multiple range test. Strontium levels in wild fish Juvenile P. auratus were caught between November 1993 and April 1994 with Opera House fish traps (100 X 80 X 60 cm, stretched mesh size 11 mm) from each of four estuarine locations around the Sydney region. From each estuary 20 fish of similar size to those used in experiments one and two were selected: those from Botany Bay had a mean fork length of 63 mm (SD ±5.9), from Port Jackson, 80 mm (SD ±9.9), from Middle Harbour, 82 mm (SD ±4.4), and from Port Hacking, 80 mm (SD ±5.6). The main aim of this experiment was to compare naturally occurring levels of dorsal spine strontium in wild-caught juvenile snapper from multiple loca- tions, of potentially different water chemistries, with 1) snapper immersed in Sr 40x ambient for 48 hours, and 2) snapper immersed in Sr lOx ambient for 5 days. Statistical analysis involved one-way analysis of variance and the SNK multiple range test. Treatment of experimental Fish The experiments were undertaken in an enclosed room containing thirty-two cylindrical 100-litre fibreglass tanks, each with a 500-|im mesh filter, an aquarium air stone and a clear plastic cover. Twice daily water exchange from Salamander Bay was pro- vided by a 1000-litre supply tank for each row of eight tanks. Photoperiod was ten hours of light per day, and tanks were cleaned weekly. Approximately 0.125 g of feed per fish per day was distributed in three equal portions throughout the day. The specially for- mulated dry snapper diet was a 50% protein extruded diet based on 649c fish meal, similar in composition to that in Quartararo et al. (1992). To maintain a constant stocking density, fish mortalities were com- pensated for by replacement with similar-size un- treated snapper, fin-clipped for identification. The water quality parameters measured every sec- ond day were pH (which averaged approximately 8.1), electrical conductivity (approx. 52 mS/cm), turbidity (generally 0 or 1 NTU), dissolved oxygen (approx. 8.6 mg/L), water temperature (approx. 19-23°C ), and salinity (approx. 34.3 ppt). No significant differences were observed between tanks for any of these pa- rameters, which remained within a range generally acceptable for the maintenance offish health ( Poxton andAllouse, 1982). Treatments were randomly allocated to tanks. All fish chosen appeared healthy and in good condition. Wet weight and fork length were measured prior to introduction of representative groups to the experi- mental tanks. Sedation was necessary to allow safe handling and was achieved with 0.5 mL/L of ben- zocaine solution, which comprised 100 g/L ethyl p- amino benzoate (CgH^jNO,) in 70% ethanol solvent (Summerfelt and Smith, 1990). The fish were accli- matized to the tanks for a period of 5 days for experi- ment 1 and 8 days for experiment 2. Strontium treat- ments were initiated by dissolving the appropriate quantity of chemical salt in each tank. For the dura- tion of the treatment there was no water exchange or cleaning activities. There was also no feeding of the fish during treatment to avoid uneven uptake of the chemical marker among individuals as a result of differential feeding (e.g. Gallahar and Kingsford, 1996). Treatment was terminated after the prescribed exposure period by allowing the exchange of new water. Fish were removed by dip net from the tanks and placed into a lethal dose of 1.0 niL/L benzocaine solution at the appropriate time of sampling, then measured, and frozen for storage. Dorsal spine tissue samples were dissected from the treated and control snapper Each sample con- sisted of four sturdy spines from the midfront of the dorsal row. The weaker posterior spines were avoided where possible. These samples were then placed into polypropylene eppendorf tubes and dried overnight in a 75°C oven. The tubes were then coaled for half Pollard et al.: Chemical marking of Pagrus auratus 121 Source an hour prior to weighing. The tubes were sealed during this time to reduce moisture uptake which could have caused an increase in weight. Samples were weighed to the nearest 0.001 mg. The remaining steps took place in a laminar flow cabinet to avoid contami- nation. With a pipette we placed 50 |iL of 1% nitric acid (H2NO3) into the ep- pendorf tube containing the spines for 10 seconds and then removed the acid. This procedure cleaned possible exter- nal contaminants from the surface of the sample. The time and concentration needed for cleaning were determined by microscopic examination of the spines to check for excessive etching and loss of bone. The contents of the eppendorf tube were then dissolved in a proportional amount of 75% nitric acid to produce a standard 1 mg of spine tissue per 10 |iL of solution. After allovraig an hour or more for the sample to fully dissolve in the sealed tube, 10 |iL of solution was removed and added to a sample tube containing 4950 |iL of milliQ water and 40 |iL of concentrated nitric acid. This resulted in a standard- ized 0.2 mg/|iL sample solution in 1% nitric acid which became the sample concentration used in our experiment after preliminary analyses with ICP-MS. Measurement of isotopes Inductively coupled plasma-mass spectrometry (ICP- MS) was used to measure concentrations of stron- tium in the dorsal spine tissue. The model used was a Perkin-Elmer SCIEX ELAN 5000 with a Gilson 212B autosampler for sample introduction, and pro- cedures were similar to those described in Dove et. al. (1996). The ICP-MS was optimized with a 0.010 mg/L standard solution covering the extremes of the mass range. By using the graphics application, we were able to adjust the nebulizer flow to maximize the i°^Rh signal and balance the signals for ^'^Mg and 208pij fj.pj^ ^^Q standard solution. The quantitative analysis function of the ICP-MS was externally cah- brated by using calibration standards with concen- trations of 1, 10, 100, and 500 parts per billion, i.e. those covering the range of expected concentrations. The correlation coefficient of the calibration curve was generally 1.000 for *^Sr, the isotope used for analysis. ICP-MS interferences such as isobaric overlaps or plasma-induced polyatomic ions are generally negligible for strontium. A blank solution containing only the 1% nitric acid matrix was also analyzed prior to the sample solutions, so that the isotopic concentration of the acid matrix could be subtracted. Table 1 Analysis of variance for experiment 1: persistence over 36 days. Three- way partially hierarchical analysis of variance to test for differences in dorsal spine *^Sr among sample times 0, 12, 24, and 36 days (fixed factor=time). treatments (treat) (Sr^* zero, 5x and lOx ambient; fixed fac- tor), and tanks (random factor). Nonsignificance at P=0.05 is indicated by n.s. Cochran's test; df = 1; no. variances (*) = 48; and C = 0.1565; variances homogeneous; no transformation. df Mean square F-ratio Probability Time 3 13699.2 6.690 <0,005 Treatment 2 568635.5 105.485 <0.001 Time X treat 6 4536.0 2.215 n.s. Tanks (treat) 9 5390.7 2.632 <0.050 Time X tanks (treat) 27 2047.8 0.695 n.s. Residual 48 2946.5 Quality control solutions were analyzed for *^Sr at the start, middle, and end of each analytical run for each date of anaiysis, allowing the stability of the anal3^e signal to be monitored both within and be- tween runs. The quality control readings obtained were not constant over time, owing perhaps to clogged sampling cones. The results for the spine samples, therefore, required adjustment in relation to the quality control results. We assumed that there was no change in the *^Sr concentration of the quality control solution over time. External drift correction can significantly improve both accuracy and preci- sion (Jarvis et al., 1992). Treatment of data Data were analyzed by analysis of variance accord- ing to the general recommendations of Underwood (1981). Normality of data was tested with Cochran's test. Student-Newman-Keuls (SNK) multiple range tests were used for a posteriori comparisons among treatments where appropriate. Where data were missing (<2 per design), the missing values were re- placed by the means for their group, and the degrees of freedom of the residual were reduced by the num- ber of missing values. Results Persistence of strontium over 36 days Differences in the ^^Sr concentrations of dorsal spines were found among treatments, despite differences among tanks within treatments (Table 1; Fig. 1). Fish exposed to strontium at lOx ambient incorporated more strontium into their spines than fish at 5x 122 Fishery Bulletin 97(1), 1999 ambient (SNK: ControkSr 5x o c c CO X CO X O Q> ° in ■c a> tr ^ r o o u. Q- Q_ Figure 3 Naturally occurring levels of "''Sr in the spines of wild Pagrus auratus from four estuaries in New South Wales, Australia, compared wirh those of hatchery-reared fish (controls from Port Stephens, N.S.W.) and fish treated with strontium. Behrens-Yamada et al. ( 1979) found that strontium appears to become metabolically inert in the verte- brae of coho salmon, Oncorhynchus kisutch, after 124 Fishery Bulletin 97(1), 1999 bo 3 a a o a x; t> !c Q. O o *^ 13.2 < s c X S a- o -^ M ii -o c m — X m Qi ffl O "O C -c "^ O -J= s 0) •— o -S-^ c OJ S S « a =*- t. K O CO *^ J2 6 g ^ t c 1 3 o e-ui .2 £t) ■^ a. V C ,. X o " w -u 03 3 CO -O M be X >> (M -^ T3 .s 7 j= c ?i. to c o o c - a) a 3 2 T3 3 o -2 -i K — s «^ M O O hr ^ OJ o CO a ^ (ft O -5 "^ T3 .-H 3 O T1 CJ Oi f/l c '/; OJ x: r: S o « £ ^ in o m i-n m ^ s-g S <» 2-S T3 O Qj CO 3 >- -a ^ a-" o t>o rH t; -a lis CQ >■ _ 3 XI 3 m ^ Oj o X 'fl cu cu 0.^3 (Tl (3 en o X: a 3 Vi Tl n 3 C3 c -? C o c S O X 3 '§"£■*: CO a O tJ ^ o -C ^ K r. p o O -~ tuj a o _ -« o T3 C f ^ CO 2 CO -; M Ills oa >■ 2 _ XI CD -J 2 3. cfi X S X ^ ^ >5 pa >< 2 T1 M u X s 60 c 0; C O -3 CO ■Si < IS C 01 ■T X en :S 0 C % S£cD § s ^ 01 X ■* s x: 2 CO 5^ O 3 c ^ O 2 ° £ •- O "■ Of s = s X I- c " S2 D- c j: J I £ a X X c CO c i-a: 2 9 E = 2 CO g y] o &■£ c — ■ - CO CTj !U S - Oi o # o CO fe ^ <3J n C/J C3 tac CO I .o S5 Q 3 a o ^ ^ 2 c X a s ■3:5 c:^ S M _; CO _: C in X c CO n cfi S 0 CO CQE IN I. 0. a CJ X >> 0 c CO 5 c m CO CO C 0 fc: c 0) .2 ti; E T3 -J X >- ■tix Cfi X ■- |x 0 CO CO CO 0 >> CM s a. 3 a Eii^ CO n a C Oj L. rrt S 2 3 K '.n c CC CO ■^ 0 X a* CO a c 2 X 0 11 x; X 3 < T3 B CO ■a rji 2 ^ CO CO Zj 00 o ^ c ■- ■- 73 ■2 "t 0; ^- S X C3 CO _ CO CO 0! 3 > CJ CO OJ J3 9- o o , .:: a c °- ■2§ CO CO S i- ■-2 3 Ex-^ £ C Tf .- CO CM -s: _ i ?; c o o SCO CO 0. o a — o o O CO 0 I. 01 3 o c j: .2 V. x ,9 |5 ■S w — CO 2 a k, t- U s -^ si -— CO o CJ Cy = 3 CO CO X ^ ^ c k. k. ■-■2 a> OJ CO 0 T3 - X -c '2 _ c ir: 0 CO CO 2i^ ■S2 en C CJ cn 05 CO .2 0^ ^ ofcc:-:: Pollard et al.: Chemical marking of Pagrus auratus 125 incorporation. Their inability to identify strontium- marked adults in following years was attributed en- tirely to dilution caused by the growth of vertebrae, rather than a process of leaching from the tissue. Behrens-Yamada and Mulligan ( 1982 ) found that the use of smaller vertebral core samples allowed the identification of strontium-labelled adult salmon. The limiting core diameter was related to the vertebral di- ameter of the juvenile salmon at the time of treatment. Similarly, the growth of juvenile snapper spines will cause a dilution of the strontium signal. The ratio of marked to unmarked spine matrix will decrease as the spine enlarges. Spine gi-owth occurs from the base upwards (e.g. McFarlane and Beamish, 1987), and the strontium mark may remain only in the up- per regions as spine elongation and enlargement oc- curs. Only the uppermost section of the spine, corre- sponding to the length of the spine at the age of in- corporation, should be sampled in large fish; this will result in a proportional increase in the amount of strontium in the sample. Papadopoulou et al. (1980) found that a range of chemical constituents in the otoliths of the mackerel Scomber japonicus colias decreased with age. Dilu- tion from growth, dietary changes with age, and otolith compositional changes were mentioned as possible causes. Nevertheless, the duration of most tagging experiments does not generally exceed a pe- riod of several years, and growth or age-related prob- lems are unlikely to be a major concern over such time scales. The differences between the strontium- tagged and wild fish are great enough for there to be little confusion between the two groups if samples are taken from the uppermost spine region and tag- ging experiments are not designed to extend over the very long term. Strontium in tagged and wild fish It was a major concern for our study that naturally occurring variation in the levels of strontium in wild fish might obscure the strontium signal in fish that had been tagged by immersion in SrCl2. Adult snap- per may undertake movements associated with feed- ing or spawning (Crossland, 1976), extending the range of water chemistries experienced by the fish. Moreover, juveniles may remain resident in shallow bays and estuaries that are subject to variable inputs of fresh water (Kingsford and Suthers, 1994) and that are likely to have differences in water chemistry. Despite the observed variation in natural levels of strontium in wild snapper, the treatments with Sr lOx for 5 days (Fig. 3) and Sr 40x for 24 to 48 hours (Figs. 2 and 3) had unequivocally higher readings. Other treatments also had higher readings. They were, however, not considered great enough for ef- fective tagging owing to the increased variation in spine strontium expected to result from growth fac- tors and exposure to the variable marine environ- ment. The minimum concentration recommended is Sr^'^ lOx ambient, with an exposure period of at least 4-5 days at this concentration. Shorter exposure periods will require concentrations greater than Sr lOx, and snapper significantly larger than those used in our study may also require higher treatment con- centrations. Although the comparisons were made with wild snapper captured within a period of six months in the general region of Sydney across a range of less than 40 km of coastline, it is unlikely that fish tagged with strontium at this recommended level would be confused with wild fish from any latitude or time. It is not surprising to find differences in naturally occurring strontium among fish from different estu- aries. For example, Edmonds et al. (1989) investi- gated eight elements, including strontium and found elemental compositions to be specific to the geo- graphical origin of nonmixing populations of snap- per in Western Australia. Similarly, Port Jackson, Botany Bay, and Port Hacking are distinct estuaries which would also be expected to contain nonmixing juvenile snapper populations. On the other hand, Middle Harbour is within Port Jackson, and inter- estingly no significant difference has been observed between these two groups. A variety of factors influence natural variation in the Sr content of otoliths. Salinity is perhaps the strongest determinant, and analysis of strontium levels has long been used to interpret the salinity histories of anadromous or catadromous fish species (Castonguay and Fitzgerald, 1982; Coutant and Chen, 1993; Limburg, 1995; Secor et al, 1995; Pender and Griffin, 1996). Decreasing water temperature causes an increase in otolith Sr/Ca ratio, with the greatest effect occurring at low temperatures. This knowledge has been used in attempts to reconstruct temperature histories of wild fish (Radtke and Targett, 1984; Radtke et al., 1990; Townsend et al., 1989, 1992, 1995, Arai et al., 1996). It has been sug- gested that low temperatures may impair physiologi- cal mechanisms that exist in fish and inhibit the uptake of strontium (Kalish 1989; Townsend et al., 1992). Other environmental and biological factors that influence the natural markings of calcified struc- tures include pH(Moreauetal., 1983;Wickins, 1984), fish age and size (Papadopoulou et al., 1980; Gauldie et al., 1995), developmental and reproductive events (Francis, 1994; Kingsford and Atkinson, 1994), physi- ology (Kalish, 1991), periods of stress (Townsend et al., 1992), dietary differences (Edmonds et al., 1989; 126 Fishery Bulletin 97(1), 1999 Limburg, 1995; Gallahar and Kingsford, 1996), and anthropogenic pollutants (Kalish, 1995). All of these factors may cause variation in levels of strontium in wild fish. Hence, if strontium tagging is to be suc- cessful in a region, the magnitude of natural varia- tion should be investigated at appropriate spatial and temporal scales. Strontium tagging has been used for freshwater and anadromous species (Table 3), but investigations of saltwater species are limited. Hurley et al. ( 1985) marked the statoliths of the short-finned squid lllex illecebrosus and Gallahar and Kingsford (1996) marked the rock blackfish Girella elevata. Because seawater has some 200-400 times the strontium con- tent of freshwater (Guillou and de la Noue, 1987), the range of concentrations available for tagging (i.e. before saturation and precipitation ) is reduced for marine fish. This limited our ability to batch mark separate groups offish with different concentrations of the same element, especially considering the un- known factors affecting long-term persistence, such as dilution due to growth. Multiple immersions in strontium have, however, created multiple, discrete bands on the otoliths of salmon, allowing the pro- duction of codes for batch marking (Schroder et al., 1996). Another attractive possibility is to mark batches of fish with different combinations of sev- eral elements. A characteristic of ICP-MS analysis is the ease in identification of a number of isotopes with little corresponding increase in preparation time. Schroder et al. ( 1996) found that rubidium was incorporated into the calcified tissues of juvenile salmon. Caesium, which has a similar ionic radius and the same +2 valence as calcium, may also prove to be a likely candidate for consideration. Pollard'^ investigated iron chloride ( FeCl3-6H,,0 ) together with strontium chloride, however the iron was not success- fully incorporated into the dorsal spines of snapper. Advantages, limitations, and applications of the technique The strontium immersion technique allows hundreds offish to be tagged in a single batch without the need for individual treatment. Other more labor intensive methods may introduce problems of stress from fish handling and are unsuitable for large numbers, es- pecially the very large numbers involved in reseed- ing programs. Stock enhancement in Japanese wa- ters exceeds 15 million juvenile snapper each year (Tsukamoto et al., 1989). Strontium tagging may be applied to very small fish and even larval fish (e.g. Behrens-Yamada and Mulligan, 1987). Fast body growth and the difficulties of attachment make the conventional tagging of small fish extremely diffi- cult. Some alternatives to batch tagging with stron- tium and with comparable advantages include microwire tags (Beukers et al., 1995), thermal mark- ing (Schroder et al., 1996), oxytetracycline (Francis et al., 1992; Lang and Buxton, 1993), alizarin complexone (Lang and Buxton, 1993; Secor and Houde, 1995) and other chemical fingerprints (e.g. Gillanders and Kingsford, 1996). It may be useful for some purposes to use a combination of techniques simultaneously. The costs associated with the analysis of samples on ICP-MS are relatively high compared with those of many conventional tagging techniques, however they may compare favourably with the costs of some other chemical, electronic, or genetic methods of iden- tification. The ICP-MS has very low detection lim- its, generally 100 to 1000 times more sensitive than inductively coupled plasma-atomic emission spectro- scopy and inductively coupled plasma-atomic fluo- rescence spectrometry (Horlick and Shao, 1992). Sample preparation for ICP-MS is labor intensive; around 100 fish samples require up to 2 days prepa- ration and a half day for analysis. Strontium marks however may also be detected by using other ana- lytical techniques with lower costs, such as energy- dispersive x-ray spectrometry or wavelength disper- sive spectrometry (Schroder et al., 1995). Many other analytical techniques, such as backscattered electron microscopy (e.g. Schroder et al., 1995), or laser abla- tion ICP-MS (e.g. Fowler et al., 1995), are also de- signed to measure microconstituents at specific loci across a section of calcified tissue, allowing the ap- plication of strontium marking for the validation of aging and batch marking of groups offish. Otoliths are commonly used for chemical marking and are the most widely used structures for age de- termination (Campana and Neilson, 1985). Otoliths comprise discrete, directly comparable samples and are recognized for their elemental stability (Edmonds et al., 1989). However, fish must be sacrificed for otolith sample collection. By contrast, spine and scale removal are nondestructive techniques, allowing catch and release techniques to be used for tagged fish. Spines are not shed or replaced during growth, unlike scales which may be lost, resorbed, or regen- erated during fish ontogeny (Coutant and Chen, 1993). The small size of juvenile snapper scales also creates counting, weighing, and manipulation difli- culties (Pollard*). The dissection of otoliths or other internal skeletal structures from large groups offish is a labor intensive activity. The ease of removal of spines or scales allows for the routine sampling of these tissues regardless of the original purpose of fish collection. Removal for sampling could take place before the distribution of fish for commercial pur- Pollard et al.: Chemical marking of Pagrus auratus 127 poses. Removal of spines or scales causes minimal physical defacement of the product, an aspect of particular relevance to the Japanese market where aesthetic appeal is highly valued. The immersion technique for incorporation of strontium enables greater control over the degree of exposure offish to strontium than does introduction through the diet. The latter method may introduce inconsistencies between experimental fish as a re- sult of differential feeding, which often results from size hierarchies within the tanks (Umino et al., 1993). Intraperitoneal or intramuscular injection allows similar control of exposure to that of immersion; how- ever each fish must be dealt with separately. Scott (1961) was unable to mark fish under 8 cm long by using injection techniques. Nevertheless, with large pelagic fish injection provides many advantages, in- cluding the ability to tag directly from the capture vessel. The strontium tag showed no signs of decay due to leaching from the spine tissue, and tag retention times may prove to compare favorably with conven- tional methods. External physical tags are shed at different rates for different species, fish sizes, tag types, or attachment sites (e.g. Ingram, 1993), and may facilitate disease. Tattoos may be rendered un- readable by fish growth, and vital dyes may be leached from the animal (Laird and Stott, 1978). Clipped fins may regenerate, or there may be confu- sion between clipped fins and fins that have been naturally excised by predators. Population studies also assume that the proportion of tagged to untagged fish captured is representative of that existing in the population. These assumptions hold for strontium marking, whereas individuals marked with tags may be less fit for survival, more conspicuous to predators, or more likely to become entangled in capture nets. Radioactive isotopes are effective markers offish, as evidenced by the permanent marking of many fish by ^^C after atomic bomb testing in the 1950s (Kalish, 1995). A number of different radioactive isotopes have been investigated for fish tagging, for example ^^Ca (Bogoiavlenskaia, 1959;Anwand, 1966), -^^PiKarzin- kin et al., 1959), ^^Fe (Scott, 1961), ^^"Cs (Scott, 1962), I'^^Ce (Hoss, 1967), ^^il (Fitzgerald and Keenleyside, 1978), is^Eu and ^^^Eu (Hansen and Fattah, 1986), ^^Sr (Carlson and Shealy, 1972; Lehtonen et al., 1992), s^Sr (Farrell and Campana, 1996), and ^^Sr (Zhao et al., 1992). Marking with stable strontium salts however has many advantages over radioactive tagging, which is comparatively expensive, environ- mentally unattractive, and a potential threat to hu- man health. The latter is an important criticism when radioisotope techniques are used to mark com- mercially important species, such as radioactive iri- dium used to tag and release snapper (Kato, 1990; Kato et al., 1991). Conversely, strontium immersion has in fact been suggested as a low-level marking technique for farm-reared brook trout destined for human consumption (Guillou and de la Noue, 1987). A major disadvantage of the technique is that the existence of the strontium mark cannot be detected in recaptured fish prior to analysis. This means that large batches of snapper of the appropriate size may have to be analyzed when only a small proportion may be marked individuals. Most conventional tag- ging methods, including most radioactive isotope markers, allow in vivo detection of tagged fish. The number of fish requiring analysis should be mini- mized by the preselection of appropriate-size fish based on growth rate estimates for the appropriate geographic location (e.g. Francis and Winstanley, 1989; Tsukamoto et al., 1989; Paul, 1992). Moreover, other external features may also aid in the preselec- tion of possibly marked fish; for example, a large proportion of hatchery-reared Japanese snapper lack an internostril epidermis (Sobajima et al., 1986). Salmon stocks of different origin have been identi- fied by using naturally occurring differences in scale patterns (Bilton, 1972), otolith patterns (Hindar and Abee-Lund, 1992), vertebral elemental composition (Mulligan et al., 1983), or parasitic fauna (Margolis, 1963). Stock enhancement or reseeding programmes may be of particular importance when there has been re- cruitment failure due to natural or anthropogenic environmental change, or as a result of overfishing. Pagrus auratus have been ranched and released for many years in Japan. Concerns have been raised about alterations to the gene pool from the introduc- tion of genetically inferior cultured fish (Hindar et al, 1991; Harada, 1992). Smith and Hataya (1982) have however estimated that the release of one mil- lion snapper per year would increase annual net re- turns to the fishery by about 35 metric tons, and the Kagoshima Bay ranching operation appears to have become highly profitable ( Matsuda, 1992 ). There has also been consideration of the establishment of snap- per reseeding operations in the southern hemisphere (Smith and Francis^). Strontium tagging can assess the economic or conservation value of such stock en- hancement operations and lead to refinement of the reseeding techniques, for example selection of the optimal locations and fish sizes for release. Consid- eration of migratory patterns (e.g. Crossland, 1976; Edmonds et al., 1989) is required to achieve spatial separation of studies, and temporal division of stud- ies may also be achieved by the use of different co- horts separated by sufficient time to allow size sepa- ration. The success of the strontium immersion tech- 128 Fishery Bulletin 97(1), 1999 nique for Pagriis auratus indicates that it is prob- ably applicable to many other saltwater fish species, and perhaps also to crustaceans and moUusks, but experimental validation would be required for each species. Conclusions The present study has demonstrated that the snap- per Pagrus auratus can be reliably tagged by the in- corporation of strontium into the dorsal spines. The use of strontium as a chemical marker can allow large numbers of small hatchery-bred fish to be tagged before release into the wild. Importantly, fish do not have to be sacrificed to assess whether they have been tagged. Strontium in the spines of snapper persisted for at least 36 days and there was no suggestion that the chemical signal decayed over this time. It is rec- ommended that juvenile snapper require immersion in treatments of strontium chloride greater than 0.25 g/L (lOx ambient) for 4 to 5 days for the production of reliable strontium marks in spines. Although natu- ral levels of strontium in spines varied between some locations along the central coast of New South Wales, this variation was not great enough to obscure the differences between tagged and wild fish. It was con- cluded, therefore, that strontium immersion is a use- ful and relatively environmentally safe method of tagging large numbers of small snapper. Acknowledgments The authors would like to thank D. Ferrell from the N.S.W. Fisheries Research Institute, P. Snitch from the Royal Prince Alfred Hospital, W. Talbot, J. Cleary and P. Beevers from the Brackish Water Fish Cul- ture Research Station, and S. Dove and B. Gillanders from the University of Sydney. For constructive com- ments on the manuscript we thank D. Pollard, S. Dove, D. Ferrell, and B. Gillanders. Financial sup- port was provided by the Australian Research Coun- cil to MJK. Literature cited Anwand, K. 1966. Die Anreicherunt; von ■* 'Ca unci ^'Y in Brut von £.s-o.t lucius L. unci Anguilla arifiuUUi L. Verhiancllungen der Internationale Vereinij^ng der Limnologie. 16:1 124-1 129. Arai, N., W. Sakamoto, and K. Maeda 1996. Correlation between ambient .seawater temperature and strontium-calcium ratios in otoliths of red sea bream Pagrus major. Fish. Sci. (Tokyo) 62:652-653. Battaglene, S., and J. Bell. 1991. Aquaculture prospects for marine fish in New South Wales. Fishnote DF/6, NSW Agriculture and Fisheries, Sydney. Australia, 3 p. Battagiene, S. C, and R. B. Talbot. 1992. Induced spawning and larval rearing of snapper. Pagrus auratus (Pisces: Sparidael. from Australian waters. N.Z. J. Mar Freshwater Res. 26:179-183. Behrens-Yamada, S., and T. J. Mulligan. 1982. Strontium marking of hatchery-reared coho salmon, Oncorhynchus kisutch Walbaum, identification of adults. J. Fish Biol. 20:.5-9. 1987. Marking nonfeeding salmonid fry with dissolved strontium. Can. .J. Fish. Aquat. Sci. 44:1.502-1.506. Behrens-Yamada, S., T. J. Mulligan, and S. J . Fairchild. 1979. Strontium marking of hatchery-reared coho salmon Oncorhynchus kisutch. Walbaum. J. Fish. Biol. 14:267-275. Beukers, J. S., G. P. Jones, and R. M. Buckey. 1995. Use of implant micro-tags for studies on populations of small reef fish. Mar. Ecol. Prog. Ser. 125:61-66. Bilton, H. T. 1972. Identification of major British Columbia and Alaska runs of even-year and odd-year pink salmon from scale characters. J. Fish. Res. Board Can. 29:295-301. Bogoiavlenskaia, M. P. 1959. A study of calcium metabolism with a view to utilising ■••'Ca as a mark for fish. Vsesoiuznyi N.-I. Institut Morskovo Rybnovo khoziaistva i Okeanografii, 55 p. Brennan. J. S., and G. M. Cailliet. 1989. Comparative age-determination techniques for white sturgeon in California. Trans. Am. Fish. Soc. 118:296- 310. Brown, P., and J. H. Harris. 1995. Strontium batch marking of golden perch iMacquaria ambigua Richardson) (Percichthyidae) and trout cod iMaccullochella macquariensis) (Cuvier). In D. H. Secor. S. E. Campana, and J. M. Dean (eds. ), Recent developments in fish otolith research, p. 693-701. Univ. South Caro- lina Press. Columbia. SC. Campana, S. E., and J. D. Neilson. 1985. Microstructure of fish otoliths. Can. J. Fish. Aquat, Sci. 42:1014-1032. Carlson, C. A., and M. H. Shealy. 1972. Marking larval largemouth bass with radiostron- tium. J. Fish. Res. Board Can. 29:45.5-459. Castonguay, M., and G. J. Fitzgerald. 1982. Examination of the method of discriminating between anadromous and freshwater fish of the same species by the content of strontium in the scales. Can. J. Fish. Aquat. Sci. 39:1424-1425. Coutant, C. C, and C. H. Chen. 1993. Strontium microstructure in scales of freshwater and estuarine striped bass iMorone saxatilis) detected by laser ablation mass spectrometry. Can. J. Fish. Aquat. Sci. .50:1318-1323. Crossland, J. 1976. Snapper tagging in north-east New Zealand, 1974: analysis of methods, return rates, and movements. N.Z. J. Mar Freshwater Res. 10:675-686. 1980. Population size and exploitation rate of snapper. Chrysophrys auratus, in the Hauraki Gulf from tagging experiments, 1975-76. N.Z. J. Mar. Freshwater Res. 14:255-261. Davy, B. 1990. Mariculture in Japan: development of an industry. World Aquacult. 21:36-47. Pollard et al ; Chemical marking of Pagrus auratus 129 1991. Mariculture in Japan: current practices. World Aquacult. 22:30-35. Dove, S. G., B. M. Gillanders, and M. J. Kingsford. 1996. An investigation of chronological differences in the deposition of trace metals in the otoliths of two temperate reef fish. J. Exp. Mar Biol. Ecol. 205:15-33. Edmonds, J. S., M. J. Moran, and N. Caputi. 1989. Trace element analysis of fish sagittae as an aid to stock identification: pink snapper (Chrysophrys auratus) in Western Australian waters. Can. J. Fish. Aquat. Sci. 46:50-54. Farrell, J., and S. E. Campana. 1996. Regulation of calcium and strontium deposition on the otoliths of juvenile tilapia, Oreochromis niloticus. Comparative Biochemistry and Physiology A 115:103-109. Ferrell, D. J., G. W. Henry, J- D. Bell, and N. Quartararo. 1992. Validation of annual marks in the otoliths of young snapper, Pagrus auratus (Sparidae). Aust. J. Mar. Fresh- water Res. 43:1051-1055. Fitzgerald, J. G., and H. A. Keenleyside. 1978. Technique for tagging small fish with 1131 for evalu- ation of predator-prey relationships. J. Fish. Res. Board Can. 35:143-145. Foscarini, R. 1988. A review: intensive farming procedure for red sea bream iPagrus major) in Japan. Aquaculture 72:191-246. Fowler, A. J., S. E. Campana, C. M. Jones, and S. R. Thorrold. 1995. Experimental assessment of the effect of tempera- ture and salinity on elemental composition of otoliths us- ing laser ablation ICPMS. Can. J. Fish. Aquat. Sci. 52:1431-1441. Francillon-Vieillot, H., V. de Buffrenil, J. Castenet, J. Geraudie, F. J. Meunier, J. Y. Sire, L. Zylberberg, and A. de Ricqles. 1990. Microstructure and mineralization of vertebrate skel- etal tissues. In J. G. Carter (ed.), Skeletal biominerali- zation: patterns, processes and evolutionary trends, vol. 1, p. 471-530. Van Nostrand Reinhold, New York. NY, Francis, M. P. 1994. Duration of larval and spawning periods in Pagrus auratus (Sparidae) determined from otolith daily increments. Environ. Biol. Fish. 39:137-152. Francis, R. I., L. J. Paul, and K. P. Mulligan. 1992. Ageing of adult snapper (Pagrus auratus) from otolith annual ring counts: validation by tagging and oxytetracy- cline injection, Aust. J. Mar. Freshwater Res. 43:1069- 1089. Francis, R. I., and R. H. Winstanley. 1989. Differences in growth rates between habitats of south- east Australian snapper iChrysophrys auratus). Aust. J. Mar Freshwater Res. 40:703-710. Fukuhara, O. 1985. Functional morphology and behaviour of early life stages of red sea bream. Bull. Jpn. Soc. Sci. Fish. 51:731- 743. Fukusho, K. 1991. Red sea bream culture in Japan, /n J. P. McVey (ed.), CRC handbook of mariculture, vol, II: finfish mariculture, p. 73-87. CRC Press, Boca Raton, FL. Gallahar, N. K., and M. J. Kingsford. 1992. Patterns of increment width and Sr/Ca ratios in otoliths ofjuvenile rock blackfish, Girella elevata. J. Fish Biol. 41:749-763. 1996. Factors influencing Sr/Ca ratios in Girella elevata: an experimental approach. J. Fish. Biology 48:174-186. Gauldie, R. W., I. F. West, and G. E. Coote. 1995. Evaluating otolith age estimates for Hoplostethus atlanticus by comparing patterns of checks, cycles in microincrement width, and cycles in strontium and cal- cium composition. Bull. Mar. Sci. 56:76-102 Gilbert, D. J. 1986. A stock reduction analysis of Bay of Plenty snapper N.Z. J. Mar Freshwater Res. 20:641-653. Gillanders, B. M., and M. J. Kingsford. 1996. Elements in otoliths may elucidate the contribution of estuarine recruitment to sustaining coastal reef popu- lations of a temperate reef fish. Mar Ecol. Prog. Ser. 141:13-20. Guillou, A., and J. de la Noue. 1987. Use of strontium as a nutritional marker for farm- reared brook trout. Prog. Fish Cult. 49:34-39. Hansen, H. J. M., and A. T. A. Fattah. 1986. Long-term tagging of elvers, AnguiV/a anguilla, with radioactive europium. J. Fish Biol. 29:535-540. Harada, Y. 1992. Genetic difference between wild and released indi- viduals and the resource enhancement effect of stocking: a theoretical analysis. Nippon Suisan Gakkaishi 58:2269- 2275. Hindar, K., and J. H. Abee-Lund. 1992. Identification of hatchery-reared and naturally pro- duced Atlantic salmon Salmo salar L. juveniles based on examination of otoliths. Aquacult. Fish. Manage. 23:235- 241. Hindar, K., N. Ryman, and F. Utter. 1991. Genetic effects of cultured fish on natural fish populations. Can. J. Fish. Aquat. Sci. 48:945-957. Horlick, G., and Y. Shao. 1992. Inductively coupled plasma-mass spectrometry for elemental analysis. In A. Montaser and D. W. Golightly (eds. ), Inductively coupled plasmas in analytical atomic spec- trometry, p. 55 1-612. VCH Publishers, New York, NY. Home, R. A. 1969. Marine chemistry. Elsevier, Amsterdam, The Neth- erlands. 568 p. Hoss, D. E. 1967. Marking of post-larval paralichthid flounders with radioactive elements. Trans. Am. Fish. Soc. 96:151-156. Hurley, G. V., P. H. Odense, R. K. O'Dor, and E. G. Dawe. 1985. Strontium labelling for verifying daily growth incre- ments in the statolith of the short-finned squid (Illex illecebrosus). Can. J. Fish. Aquat. Sci. 42:380-383. Ingram, B.A. 1993. Evaluation of coded wire tags for marking fingerling golden perch, Macquaria ambigua (Percichthyidae), and silver perch, Bidyanus bidyanus (Teraponidae). Aust. J. Mar Freshwater Res. 44:817-824. Ishibasi, K. 1986. A statistical assessment on the efl'ect of liberation of red sea bream Pagrus major larvae in sea-farming. Bull. Tokai Reg. Fish. Res. Lab. 119:93-107. Jarvis, K. E., A. L. Gray, and R. S. Houk . 1992. Handbook of Inductively Coupled Plasma Mass Spectrometry. Blackie and Son, London, U.K., 380 p. Jones, G. K. 1981. Biological research on snapper (C/irysop/jrvs auratus, syn. unicolor) and an analysis of the fishery in Northern Spencer Gulf SAFIC (South Australian Department of Fisheries, Adelaide) 5(6):5-8. Kalish, J. M. 1989. Otolith microchemistry: validation of the effects of 130 Fishery Bulletin 97(1), 1999 physiology, age and environment on otolith composi- tion. J. Exp. Mar. Biol. Ecol. 132;151-178. 1990. Use of otolith microchemistry to distinguish the prog- eny of sympatric anadromous and non-anadromous salmonids. Fish. Bull. 88:657-666. 1991. Determinants of otolith chemistry; seasonal varia- tion in the composition of blood plasma, endolymph and otoliths of bearded rock cod Pseudophycis barbatus. Mar Ecol. Prog. Ser. 74:137-159. 1995. Application of the bomb radiocarbon chronometer to the validation of redfish Centroberyx affinis age. Can. J. Fish. Aquat. Sci. 52:1399-1405 Karzinkin, G. S., E. V. Soldatova, and I. A. Shekhanova. 1959. Some results of mass marking of "non-standard" stur- geon fingerlings by means of radioactive phosphorus. Animal Migrations 1:27-40. Kato, M. 1990. A study on exploitation of iridium marking for red sea bream. Bull. Nat. Res. Inst. Far Seas Fish. 27:11-28. Kato, M., H. Sudo, M. Azeta, and Y. Matsumiya. 1991. Field experiments on iridium marking for red sea bream in Shijiki Bay. Bull. Nat. Res. Inst. Far Seas Fish. 28:21-45. Kingsford, M. J., and M. H. Atkinson. 1994. Increments in otoliths and scales: how they relate to the age and early development of reared and wild larval and juvenile snapper Pagrus auratus (Sparidae). Aust. J. Mar Freshwater Res. 45:1007-21. Kingsford, M. J., and I. M. Suthers. 1994. Dynamic estuarine plumes and fronts: importance to small fish and plankton in coastal waters of NSW, Aus- tralia. Continental Shelf Res. 14:655-672. Laird, L. M., and B. Stott. 1978. Marking and tagging. /?; T. Bagenal (ed.). Methods for assessment offish production in fresh waters, 3'''' ed., p. 84-100. Blackwell, Oxford, U.K. Lang, J. B., and C. D. Buxton. 1993. Validation of age estimates in sparid fish using fluo- rochrome marking. .S. Afr J. Mar. Sci. 13:195-203 Lapi, L. A., and T. J. Mulligan. 1981. Salmon stock identification using a microanalytic technique to measure elements present in the freshwater growth region of scales. Can. -J. Fish. Aquat. Sci. 38:744- 751. Lehtonen, H., K. Nyberg, P. J. Vuorinen, and A. Leskela. 1992. Radioactive strontium ("^'Sr) in marking whitefish [Coregonus lavaretus (L.)] larvae and the dispersal of lar- vae from river to sea. J. Fish Biol. 41:417-423. Limburg, K. E. 1995. Otolith strontium traces environmental history of subyearling American shadA/osa sapidissima. Mar Ecol. Prog. Ser. 119:25-35. Lopez, N.A. 1986. Morphohistological study of the early development stages of the red sea bream, Pagrun major. In J. L. Maclean, L. B. Dizon, and L. V. Hosillos (eds. ), First Asian fisheries forum, p. 179-184. Asian Fisheries Society, Manila, Philippines. Margolis, L. 1963. Parasites as indicators of the geographical origin of sockeye salmon, Oncorhynchu^ nerka (Walbaum), occur- ring in the North Pacific Ocean and adjacent seas. Int. North Pac. Fish. Comm. Bull. 11 Matsuda, Y. 1992. Sea bream ranching in Japan: the Kagoshima experience. Infofish Internat. 5:41-44. McFarlane, G. A., and R. J. Beamish. 1987. Validation of the dorsal spine method of age deter- mination for spiny dogfish. In L. C. Summerfelt and G. E. Hall (eds.). Age and growth of fishes, p. 287-300. Iowa State University Press, Ames, lA. Moran, M. 1987. Tagging confirms separate stocks of snapper in Shark Bay region. FINS (West Australian Department of Fish- eries, Perth) 20:.3-8. Moreau, G., C. Barbeau, J. J. Frenette, J. Saint-Onge, and M. Simoneau 1983. Zinc, manganese, and strontium in opercula and scales of brook trout iSalvetinus fontinalis) as indicators of lake acidification. Can. J. Fish. Aquat. Sci. 40:168.5-91. Mugiya, Y. and C. Satoh. 1995. Strontium-calcium ratios change corresponding to microincrements in otoliths of the goldfish Carassius auratus. Fish. Sci. (Tokyo) 61:361-362. Mugiya, Y. and S. Tanaka. 1995. Incorporation of water-borne strontium into otoliths and its turnover in the goldfish Carassius auratus: effects of strontium concentrations, temperature and 17-beta- estradiol. Fish. Sci. (Tokyo) 61:29-35. Mulligan, T. J., L. Lapi, R. Kieser, S. Behrens-Yamada, and D. L. Duewer 1983. Salmon stock identification based on elemental com- position of vertebrae. Can. J. Fish. Aquat. Sci. 40:215- 229. Ophel, L L., and J. M. Judd 1968. Marking fish with stable strontium. J. Fish. Res. Board Can. 25:1333-1337. O'Sullivan, D. 1992. Potential seen for snapper culture. Austasia Aquacuh. 6:36-37. Pankhurst, P. M., J. C. Montgomery, and N. W. Pankhurst. 1991. Growth, development and behaviour of artificially reared larval Pagrus auratus (Bloch and Schneider 1801 )( Spandae ). Aust. J. Mar Freshwater Res. 42:391-398. Papadopoulou, C, G. D. Kanias, and E. Moraitopoulou- Kassimati. 1980. Trace element content in fish otoliths in relation to age and size. Mar. Poll. Bull. 11:68-71. Paul, L.J. 1982. Snapper decline will hurt. Catch (N.Z.), Oct. :23-24. 1992. Age and growth studies of New Zealand marine fishes, 1921-90: a review and bibliography. Aust. J. Mar. Fresh- water Res. 43:879-912. Paulin, C. D. 1990. Pagrus auratus, a new combination for the species known as "snapper" in Australasian waters (Pisces: Sparidae). N.Z. J. Mar. Freshwater Res. 24:259-265. Pender, P. J., and R. K. Griffin. 1996. Habitat history of barramundi Lates calcarifer in a north Australian river system based on barium and stron- tium levels in scales. Trans. Am. Fish. Soc. 125:679-689. Poxton, M. G., and S. B. Allouse. 1982. Water quality criteria for marine fisheries. Aqua- cult. Eng. 1:153-191. Quartararo, N., G. L. Allan, and J. D. Bell. 1992. Fish meal substitution in a diet for Australian snap- per, Pagrus auratus. Proceedings of the Aquaculture Nutrition Workshop, Appendix 2:125-126. Radtke, R. L., and T. E. Targett. 1984. Rhythmic structural and chemical patterns in otoliths of the Antarctic fish Notolhenia larseni: their application to age determination. Polar Biol. 3:203-210. Pollard et al : Chemical marking of Pagrus auratus 131 Radtke, R. L., D. W. Townsend, S. D. Folsom, and M. A. Morrison. 1990. Strontiumxalcium concentration ratios in otoliths of herring larvae as indicators of environmental histories. Environ. Biol. Fish 27:51-61. Sakamoto, T. 1984. Distribution and movement of the red sea bream in the waters of the southern part of Wakayama Prefecture, adjacent to the Kii Strait, ascertained by tagging experi- ments in 1981. Bull. Jpn Soc. Sci. Fish. 50:1835-1842. Sanders, M. J. 1974. Tagging indicates at least two stocks of snapper Chrysophrys auratus in south-east Australian waters. N.Z. J. Freshwater Res. 8:371-374. Sanders, M. J., and D. G. M. Powell 1979. Comparison of growth rates of the two stocks of snap- per (C/irvsopyirvs auratus) in south-east Australian waters using capture-recapture data. N.Z. J. Mar. Freshwater Res. 13:279-284. Schroder, S. L., C. M. Knudsen, and E. C. Volk. 1995. Marking salmon fry with strontium chloride solutions. Can. J. Fish Aquat. Sci. 52:1141-1149 Schroder, S. L., E. C. Volk, C. M. Knudsen, and J. J. Grimm. 1996. Marking embryonic and newly emerged salmonids by thermal events and rapid immersion in alkaline earth salts. Bull. Nat. Res. In.st. Aquacult. 0 Isuppl. 2):79-83. Scott, D. P. 1961. Radioactive iron as a fish mark. J. Fish. Res. Board Can. 18:383-391. 1962. Radioactive caesium as a fish and lamprey mark. J. Fi.sh. Res. Board Can. 19:149-157. Secor, D. H., A. Henderson-Arzapalo, and P. M. Piccoli. 1995. Can otolith microchemistry chart patterns of migra- tion and habitat utilization in anadromous fishes? J. Exp. Mar. Biol. Ecol. 192:15-33. Secor, D. H., and E. D. Houde. 1995. Larval mark-release experiments: potential for re- search on dynamics and recruitment in fish stocks. In D. H. Secor. J. M. Dean, and S. E. Campana (eds). Recent developments in fish otolith research, p. 423-444. Univ. South Carolina Press, Columbia, SC. Smith, P. J., and M. Hataya. 1982. Larval rearing and reseeding of red sea bream Chrysophrys major in Japan. Fisheries Research Division Occasional Publication No. 39. Shizuoka-ken 415, Japan. 19 p. Snyder, R. J., B. A. McKeown, K. Colbow, and R. Brown. 1992. Use of dissolved strontium in scale marking of juve- nile salmonids: effects of concentration and exposure time. Can. J. Fish. Aquat. Sci. 49:780-782. Sobajima, N., M. Munekiyo, and H. Funata. 1986. Possibility of differentiation between the artificially- released and the wild red sea bream by means of the lack of the inter-nostril epidermis. Kyoto Furitsu Kaiyo Senta. 10:35-40. Summerfelt, R. C, and L. S. Smith. 1990. Anesthesia, surgery, and related techniques. In L. B. Schreck and P. B. Moyle (eds. I, Methods for fish biology, p. 213-272. Am. Fish. Soc, Bethesda, MD. Townsend, D. W., R. L. Radtke, S. Corwin, and D. A. Libby. 1992. Strontium:calcium ratios injuvenile Atlantic herring Clupea harengus L. otoliths as a function of water temperature. J. Exp. Mar Biol. Ecol. 160:131-140. Townsend, D. W., R. L. Radtke, D. P. Malone, and J. P. Wallinga. 1995. Use of otolith strontium:calcium ratios for hind- casting larval cod Gadus morhua distributions relative to water masses on Georges Bank. Mar Ecol. Progr. Ser. 119:37^4 Townsend, D. W., R. L. Radtke, M. A. Morrison, and S. D. Folsom. 1989. Recruitment implications of larval herring overwin- tering distributions in the Gulf of Maine, inferred using a new otolith technique. Mar Ecol. Prog. Ser. 55:1-13. Tsukamoto, K., H. Kuwada, J. Hirokawa, M. Oya, S. Sekiya, H. Fujimoto, and K. Imaizumi 1989. Size-dependent mortality of red sea bream, Pagrus major, juveniles released with fiuorescent otolith-tags in News Bay Japan. J. Fish Biol. 35 (suppl. A):59-69. Umino, T., M. Otsu, M. Takaba, and H. Nakagawa. 1993. Some characteristics of runty fish appearing in seed production of red sea bream. Nippon Suisan Gakkaishi 59:925-928. Underwood, A. J. 1981. Techniques of analysis of variance in experimental marine biology and ecology. Oceanogr. Mar Biol. Ann. Rev. 19:513-605. Welch, T. J., M. J. van den Avyle, R. K. Betsill, and E. M. Driebe. 1993. Precision and relative accuracy of striped bass age estimates from otoliths, scales and anal fin rays and spines. N. Am. J. Fish. Manage. 13:616-620. Wickins, J. F. 1984. The effect of reduced pH on carapace calcium, stron- tium and magnesium levels in rapidly growing prawns iPenaeus monodon Fabricius). Aquaculture 41:49-60. Zhao, W., S. Xu, and L. Hou. 1992. Study of absorption and accumulation of super (90) strontium in carp. China Environ. Sci. Zhongguo Huanjing 12:360-364 132 Abstract. — A commercially valuable trap fishery for spiny lobster (Panulirus margmatus) has existed in the North- western Hawaiian Islands since the late 1970s. Fisheries landings and re- search trapping show that spawning biomass and recruitment to the fishery collapsed in 1990 in the northern por- tion of the fishing ground and that there has been no recovery to the present, although recruitment re- mained strong at banks 670 km to the southeast. An advection-diffusion model is used to investigate larval transport dynamics between these two regions. The movement model is driven by geostrophic currents computed ev- ery 10 days from sea surface height obtained from TOPEX-POSEIDON sat- ellite altimetry. The larval transport simulations indicate that even though larvae have a pelagic period of 12 months, banks differ substantially in the proportion of larvae they retain from resident spawners as well as the proportion of larvae they receive from other banks. In particular, recruitment to the northern portion of the fishing grounds is weak due to a very low local spawning biomass and a very limited contribution of lar\'ae from the area of strong recruitment and high spawning biomass in the southeast. The results also suggest that satellite altimetry can provide useful information on physical dynamics for recruitment studies. Application of TOPEX-POSEIDON satellite altimetry to simulate transport dynamics of larvae of spiny lobster, Panulirus margmatus, in the Northwestern Hawaiian Islands, 1993-1996 Jeffrey J. Polovina Pierre Kleiber Donald R. Kobayashi Honolulu Laboratory, Southwest Fisheries Science Center National Marine Fisheries Service, NOAA 2570 Dole Street, Honolulu, Hawaii 96822-2396 E-mail address (for J J Polovina) Jeffrey,Polovina@noaa.gov Manuscript accepted 18 March 1998. Fish. Bull. 97:132-143 (1999). The spiny lobster. Pan ulirus margin - atus, is endemic to the Hawaiian Archipelago and Johnston Atoll. The species is found throughout the archipelago and is the target of a trap fishery in the northwestern portion of the archipelago known as the Northwestern Hawaiian Islands (NWHIl. From the early 1980s to 1990, the majority of fishery catches came from two banks, Necker Is- land and Maro Reef, located 670 km northwest of Necker Island ( Fig. 1 ). Catches during this period averaged about 60^^ from Maro Reef, 40^7^ from Necker Island. However, in 1990 there was a dramatic collapse in recruitment of 3-year-old lobsters to the fishery at Maro Reef, and other banks north of Maro; this col- lapse has been attributed to cli- mate-induced change in productiv- ity that has impacted various other trophic levels, such as sea birds, monk seals, and reef fishes (Polo- vina et al., 1994). After the recruit- ment collapse, the fishery reduced the spawning biomass to very low levels at Maro and at other north- ern banks, and then fishing ceased in these areas. Even with the ab- sence of fishing at Maro for at least six years, there is still no evidence of a recovery in recruitment as in- dicated from a time series of the relative abundance of 3-year-olds obtained from a standardized re- search survey (Fig. 2). However, at Necker, 670 nmi to the southeast, the recruitment drop at the end of the 1980s was much less severe and recovery has occurred in recent years and has supported a fishery (Fig. 2). The striking differences in recruitment levels over the past seven years between the two banks raises the possibility that there is limited larval mixing between the two banks. Current management for the lob- ster fishery is based on the hypoth- esis that recruitment to the NWHI banks comes from a well-mixed pool of larvae with contributions from the entire archipelago, and this pool of larvae oscillates seasonally along the archipelago, pushed northwest in the spring and summer with tradewind-driven Ekman transport and pushed back southeast in the fall and winter with westerly wind- driven Ekman transport (MacDon- ald, 1986). A genetic analysis during 1978-80 examined allozyme varia- tion in spiny lobsters from seven banks covering a substantial spatial range of the Hawaiian Aixhipelago and found no evidence of genetic differentiation between banks (Shak- lee and Samollow, 1984). However, a subsequent study in 1987, using a Polovina et al.: Application of satellite altimetry to simulate transport dynamics of Panulirus marginatus 133 Figure 1 The Hawaiian Archipelago. large sample size and comparing just Maro and Necker, found a statistically signifi- cant difference in allele frequency between Necker and Maro for one of the seven loci analyzed, raising the possibility that lar- val transport between the two banks may be limited (Seeb et. al., 1990). The importance of Ekman transport in larval dynamics is not known, but larval sampling has found that a significant lar- val density exists well below the shallow Ekman layer and, in particular, that lar- vae appear to make diurnal movements from about 80-100 m in the day to 10-20 m at night (Polovina and Moffitt, 1995). Thus larval transport may be more influ- enced by transport of the mixed layer or geostrophic transport than by Ekman transport. The importance of geostrophic transport is also supported by correlations between sea-level height from tide gauges and spatial patterns in the subsequent fishery catches (Polovina and Mitchum, 1992, 1994). Indeed, the flat, leaf-like shape of spiny lobster larvae suggests they are adapted for passive horizontal transport assisted by vertical migration within the mixed layer (Lipcius and Cobb, 1994). The Hawaiian Archipelago lies near the center of the Sub- tropical Gyre; therefore the geostrophic transport 2 1 1 1 1 1 1 1 1 1 1.8 1.6 K, — — — Maro 1.4 ^ 1.2 0. ^ 1. 9 1 1 0.8 ./ ^ \ \ " 0.6 \ ^^^—-..y^ ^* ' 0.4 0.2 • 86 87 88 89 90 91 92 93 94 95 96 Year Figure 2 Standardized catch rates of age-3 spiny lobster (number of lobster per trap) from summer research cruises at Maro and Necker (data from Honolulu Laboratory, SWFSC, NMFS). No cruise conducted in 1989. consists of weak flow from northwest to southeast along the archipelago, turning to the east at the southern end of the archipelago. Although the mean geostrophic transport is weak, geostrophic mesoscale features, particularly eddies, are prevalent and are likely to be significant for larval spatial djTiamics. 134 Fishery Bulletin 97(1), 1999 Since late 1992, data have been collected by TOPEX-POSEIDON (T-P) satellite altimetry and are available in near-real time. The satellite covers the ocean over a 10-day period, providing a temporal reso- lution of 10 days; however, the spatial coverage is along narrow tracks; therefore interpolation between track is required for full spatial coverage. We used this altimetry data to estimate geostrophic current and ultimately to drive a simulation model of the transport of spiny lobster larvae released from se- lected banks in order to describe the spatial and sea- sonal dynamics of larval transport. In particular, in the NWHI, the spiny lobsters spawn in the summer, during May-August, and in the winter, during No- vember and December, and the larvae are estimated to have a pelagic duration of 12 months (Polovina and Moffitt, 1995). We investigated aspects of the spatial and temporal dynamics of spiny lobster lar- val transport by simulating the movement of larvae from several representative banks from summer and winter spawning over the 12-month larval duration. mental Satellite, Data and Information Service (NESDIS), National Oceanic and Atmospheric Ad- ministration (NOAA), Department of Commerce. This project was a joint endeavor with the Naval Oceanographic Office (NOO) and the Jet Propulsion Laboratory (JPL). The altimetry data were first transmitted from the satellite to JPL, where the or- bit position was finalized with the aid of GPS track- ing data. NOO scientists performed an initial adjust- ment to the altimetry data using the JPL orbit infor- mation, then the data were forwarded to NOAA where a final orbit-related adjustment was made after the complete 10-day cycle of data was received. The data were aggregated into 1-degree latitude in- tervals along the satellite track (Fig. 3 1 and were expressed as a deviation from a 3-year (1993-95) mean altimetry. These data can be downloaded from an anonymous file transfer protocol (ftp) at falcon.grdl.noaa.gov in/pub/topex_real_time or from an internet webpage at Methods Topex altimetry data The altimetry data used in this study were obtained from a 2-day Delayed Altimeter Data project of the Laboratory for Satellite Altimetry, National Environ- http://ibis.grdl.noaa.gov/SAT/near_rt/ topex_2day.html. The data were ready for public dissemination ap- proximately 2 days after each 10-day satellite cycle was completed. Altimetry data that were not aggre- gated but were continuous along track lines are avail- 180' 177'W 174'W 171 'W 168'W 165"W 162'W 159"W 156'W 153'W 150"W 147'W Figure 3 Spatial grid of near-real time TOPEX-POSEIDON data from a 10-day cycle. Polovina et al.: Application of satellite altimetry to simulate transport dynamics of Panulirus marglnatus 135 able on CD-ROM from JPL. However, a considerable time lag may be associated with releases of these data. After the data were downloaded, a unix shell script was used to process the individual cycle data. This script uses various subroutines of the Generic Map- ping Tools (GMT) software package (the GMT soft- ware package is available from an anonymous ftp at kiawe.soest.hawaii.edu in /pub/gmt). The GMT script first smoothed and then interpolated the data using a subroutine called "nearneighbor." This subroutine placed the irregularly spaced track data onto a uni- formly spaced latitude-longitude matrix, assigning values to each latitude-longitude location (node). Each node was assigned a weighted mean of all data points within a user-specified search radius (SR); i.e. a mean of all data within a geographic circle about each node. The weighting factor was a function of radial distance from the node such that iv(r) =1/(1-1- (3r/SR)'^), hence wir) ranged nonlinearly from 1 at the node (r=0) to 0.1 at SR units from the node (r=SR). When all orbital passes of the satellite cycle were complete, a SR value of 3 degrees could adequately interpolate regions between passes, in- dependent of the relative date of the pass within the 10-day cycle, resulting in a complete grid file. For cycles with missing passes, the SR was able to be increased accordingly The SR could also be speci- fied as an absolute distance to ensure that the smoothing region was circular and not ovoid. After the deviation data were assembled on a grid, the script then added these data to another grid of equal dimensions containing the Levitus long-term mean dynamic height at 1000 m, which we assumed was representative of the mean dynamic topography for 1993-96. The Levitus data originated from the 1994 NODC World Ocean Atlas (CD-ROM data set), U.S. Department of Commerce, NOAA, National Environ- mental Satellite, Data, and Information Service (NESDIS). The grid file of absolute altimetry values (2) was then evaluated by the subroutine grid- gradient, which calculates the east-west and north- south gradients, dz/dx and dz/dy, where x and y and z are all in centimeters. These gradients were used for calculating the u and v components of the geo- strophic current as follows: and u = -(glf) v = {glf) dz^ dy dz (1) dx Q = 7.29 X 10"^ radians per second (earth angular rotational velocity); and (]) = latitude. These u and v values represent estimates at the sur- face. They can also serve as estimates of geostrophic current over the top 100-m mixed layer if integrals over the mixed layer of the horizontal density gradi- ents are negligible. For the region of interest in our study, given the weak horizontal density gradients, the error in using these surface estimates as mixed layer estimates, according to calculations with the Levitus density field, was less than 2 cm/s. This is negligible given the time and space scales we were considering. The ;/ and v values were then output to an ASCII file at 0.5-degree resolution. A separate file for each 10-day period (« = 169 from Oct 92 to Jun 97) was constructed for the entire Topex-Poseidon data set. The GMT script optionally produced a contour map of dynamic height with an overlay of geostrophic current vectors for each cycle. The resultant maps showed eddy and meander features that were likely important in recruitment processes (Fig. 4). Several authors have found excellent agreement be- tween currents estimated ft-om T-P data and ground- truthing. Comparisons between the current estimated from satellite-tracked drogued buoys and geostrophic current estimated fi-om T-P data in the western Pacific resulted in correlations of 0.924 for zonal velocity (u) and 0.760 for meridional velocity (v) (Yu et al., 1995). Comparisons between current speed from an acoustic Doppler current profiler along T-P track lines in the Hawaiian Archipelago agreed with estimates from T-P to within a few cm/sec (Mitchum, 1996). Sea surface height from T-P and a large set of island tide gauges, including those in the Hawaiian Archipelago, yielded a correlation of 0.66 (Mitchum, 1994). Movement model Individual larvae were tracked for a series of time steps starting from a given location by iteratively applying advective displacements due to water flow with additional random displacements caused by dif- fusion. The position of each larva was updated at each time step as follows: where g = 980 cm per second f = 2 Q sin (]); •^f+A/ =-^f +["(.v,,v,,n'^^ + f^/'OA<]/cos(y, (2) 2. where t = time in days; 136 Fishery Bulletin 97(1), 1999 30'N 28'N 26'N 24'N 22'N 20-N 30'N 28"N 26"N 24'N - n 22"N 20'N 180' izyw 174*W 17rW 168'W 165"W 162'W 159'W 156'W Figure 4 Sea surface height and geostrophic current estimateci from a 10-day TOPEX/POSEIDON cycle, 16-26 August 1993. Only currents 8 cm/sec or larger are plotted with arrows. Solid circles from lower right (east) to upper left (west), denote Oahu, Necker, Maro, and Midway Islands, respectively, X and V = the larval position in degrees of longi- tude and latitude respectively; u and I' = longitudinal and latitudinal compo- nents of velocity in degrees/day; e = a normal (mean 0, standard deviation 1) random variate; and D - the eddy diffusion rate in units of degree"/ day ( 1 m-/sec = 7 x 10"*' degrees/day). The first term in brackets in each equation is advec- tive displacement and the second term, diffusional displacement. The cosine function in the longitudi- nal movement equation corrects for the fact that dis- tance per degree of longitude decreases from the equator to the poles. Values of i/ and v at particular locations and times were obtained by linear interpolation from the precalculated 0.5 degree by 10-day grid of;/ and v val- ues computed from the T-P data. The time step was set to one day (A^=l) with 365 iterations for one year of simulation. The 1-day time step was used to match the spatial resolution of the velocity field, but simulations showed that the choice of time step was robust up to about two weeks. Larvae dispersed in the model be- cause a different diffusional displacement (a different e) was chosen for each larva at each time step. Once larvae were even slightly separated, dispersal was fur- ther advanced because the larvae experienced differ- ent advective displacements as u and i' varied spatially. Simulations Although lobsters presumably spawn on all banks in the Hawaiian Archipelago, for simplicity, the simu- lations released larvae from only four banks: Mid- way and Oahu (chosen as representative of the north- western and southeastern ends of the archipelago) and Necker and Maro (representative of the main fishing banks [Fig. 1] ). Each simulation released 5000 larvae at the beginning of the month from each of the four banks for the four months of the summer spawning season and the two months of the winter spawning season from 1993 to 1995. The simulations tracked the positions of all released larvae for 365 days; no larval mortality was assumed. For example, Figure 5 shows the distribution of simulated larvae 180 and 365 days after release at Maro Reef in July 1995. The distribution showed patchiness caused by underlying oceanographic features and expanded spatially over time. This simulation was based on an eddy diffusion rate of 1000 m"/sec, and the same simulation, but with the diffusion rate reduced to 100 m-/sec, showed a striking increase in patchiness and the same general dispersal (Fig. 6). The eddy diffusion rate for the NWHI is not known, but a range from 100 m-/sec to 1000 m-/sec has been proposed for another large archipelago, the Great Barrier Reef (Gabric and Parslow, 1994). Because the level of dif- fusion rate impacts larval patchiness and because we had a measure of actual larvae patchiness from numerous larval tows throughout the archipelago (Polovina and Moffitt, 1995), we decided the appro- priate level for eddy diffusion rate by comparing the frequency distribution of larvae sampled in actual lai-val tows with the frequency distribution of larvae from line transects and the simulated spatial pat- terns for different eddy diffusion-rate values. The frequency distribution from transects with an eddy Polovina et al.; Application of satellite altimetry to simulate transport dynamics of Panuliivs marginatus 137 35'N A diff.=1000m2/sec, 180 days 35'N 30'N 25-N 20"N 15'N 10'N % of total • > 0 - 0.05 0,05-0.25 0.25-05 ... 0.5 ■ 1 O >1 165"E 170'E 175"E 180' l75W 170'W 165"W 160'W 155'W 150"W 145"W 30'N 25"N 20"N 15'N 10N 35'N B diff.=1000fn2/sec, 365 days 35'N 30'N 25'N 20"N 15'N 10"N % of total ■ >0-0.05 = 0.05 - 0.25 o 0.25-0.5 O 0.5-1 O >i 165'E 170"E 175'E ' ^ .^V!^^ ;^* ^4.. J. " p ^:. jv, .' • r'.L'^.*>'^H'isa* 30'N 25'N 20'N 15'N 10'N 180' 175'W 170'W 165'W 160'W 155"W 150'W 145'W Figure 5 Simulated spatial distribution of 5000 larvae (A) 180 and (B) 365 days after release on 1 July 1995 at Maro with an eddy diffusion rate of 1000 m^/sec. Solid circles denote Oahu. Necker, and Midway Islands, and the star marks Maro Island. diffusion rate of 1000 m^/sec had a lognormal distri- bution but underrepresented the heavy tails of the actual larval-tow frequency distribution. The fre- quency distribution from transects with an eddy dif- fusion rate of 100 m"/sec had extremely heavy tails, in fact bimodal distribution. Either very few larvae were encountered or many larvae were encountered, resulting in a distribution that had too many large clumps compared with the observed distribution. However, an eddy diffusion rate of 500 m^/sec pro- duced a larval frequency distribution that closely matched the distribution from the larval tows, and this value was used in all subsequent simulations. At the end of their larval period, spiny lobster lar- vae metamorphose into pueruli that, at least for other spiny lobster species, are capable of directed hori- zontal swimming over 40-60 km (Pearce and Phil- lips, 1994). For P. marginatus, it is not known how close a larva needs to be to a bank at the end of its larval period to recruit to that bank. The 200-m iso- bath around the islands and banks of the archipe- lago is generally circular and marks the points at 138 Fishery Bulletin 97(1), 1999 35"N A diff.=100m2/sec, 180 days 35'N 30-N 25'N 20"N 15'N % of total >0-0.05 ° 0.05 ■ 0.25 I o 0.25-0.5 10"n| O 0.5-1 O >i 30'N 25"N 20'N 15"N 10"N 165"E 170"E 175"E 180" 175'W 170'W 165'W 160"W 155"W 150W 145"W 35"N B diff.=100m2/sec, 365 days 35"N 30'N 25"N 20'N 15*N 10'N % of total ■ > 0 - 0.05 - 0.05 - 0.25 o 0.25-0.5 O 0.5-1 •.. -■(*:„, ., 30'N 25'N 20'N 15"N 10'N O >1 165'E 170'E 175'E 180' 175'W 170'W 165'W 160'W 155'W 150'W 145'W Figure 6 Simulated spatial distribution of .5000 larvae (A) 180 and iB) 360 days after release on 1 July 995 at Maro with an eddy diffusion rate of 100 ni-/sec. Solid circles denote Oahu, Necker, and Midway Islands, and the star marks Maro Island. which the topography drops steeply to midocean depths. For Necker, Maro, and Oahu, the radius of the 200-m isobath is about 70 km. Oceanographic features resulting from interactions with topography that larvae might detect will certainly extend some distance beyond the 200-m isobath, perhaps a dis- tance equal to the radius of the bank. Thus, as an index of larval recruitment to a bank, we assumed that larvae that were within 140 km of a bank 365 days after release would recruit to that bank. The choice of 140 km was somewhat arbitrary but be- cause we used the index as a relative and not abso- lute index, the results were not particularly sensi- tive to this distance. Results Simulations were performed to describe the lai-val spa- tial dynamics from 5000 lai-vae released from all four banks for the 4-month summer spawning season ( May- August) and the 2-month winter spawning season (No- vember and December). The results indicated that there are spatially distinct patterns that generally persist Polovma et al.: Application of satellite altimetry to simulate transport dynamics of Panulirus marginatus 139 between years and spawning seasons and that were seen, for example, in the distribution of 1-year-old lar- vae in July 1995 (Fig. 7). Larvae released from Mid- way were advected eastward, larvae from Maro were advected east and south, larvae from Necker were ad- vected east and south, and larvae from Oahu were ad- vected primarily southwest ( Fig. 7). A fairly persistent meander at about 26°N concentrated larvae zonally. whereas persistent westward flow at the southern end of the archipelago advected Oahu larvae westward (Fig. 7 ). At the center of the archipelago, specifically at Maro and Necker, a substantial portion of larvae remained close to the bank where they were spawned, even afl;er 12 months, whereas at the ends of the archipelago, at Midway and Oahu, advection was high (Fig. 7). Fur- ther, although Necker receives abundant larvae spawned at Maro, relatively few larvae from Necker reached Maro (Fig. 7). The results, averaged over three years, showed that the percentage of larvae released at a bank and that were within 140 km of that bank 365 days later varied considerably by bank (Table 1). Differences in this percentage between winter and sum- mer spawning seasons were not generally large but the interannual range of this percentage often varied more than threefold (Table 1). Averaged over all three years and over the two spawning seasons, the percent of lar- vae within 140 km of a bank 365 days after being re- leased at that bank was 5.5'7r, 9.6'7f , 16.7'7f , and 16.6% for Midway, Maro, Necker, and Oahu, respectively (Table 1). From the three years of simulations, no radical changes from the bank-specific larval distribution patterns shown in Figure 7 were seen, only spatial extensions or contractions. For example, the distri- bution of larvae 365 days after spawning at Necker had a northeast and southwest orientation, and lar- vae spawned in 1993 at Necker Island were advected east and southwest to a greater extent than larvae in 1994 or 1995 (Fig. 8). Table 1 Mean percentage of larvae that were within 140 km of their | bank of origin 365 days after release 1993-96. Release Mean percent season Bank (interannual range) Summer Midway 3.1(1.8-4.7) Maro 10.4(3.9-20.4) Necker 15.7(9.5-20.3) Oahu 11.7(8.4-15.9) Winter Midway 7.9(3.7-15.0) Maro 8.8 (4.9-13.8) Necker 18.0(8.9-29.4) Oahu 21.4(12.1-33.3) To examine the exchange of larvae between banks, the recruitment index was calculated on the basis of the bank at which the larvae originated, for both sum- mer and winter spawning seasons, with 5000 larvae released from each bank (Table 2). The results were similar for both spawning seasons. Recruitment to Midway came almost entirely from larvae spawned at Midway. Recruitment to Maro was based largely on larvae from Maro and Midway; only 10-21% of the re- cruits at Maro came from Necker and Oahu. Necker had the broadest geographic source of recruits, receiv- ing substantial larvae from all four banks. Oahu's re- cruits come from Maro, Necker, and Oahu (Table 2). Discussion The altimetry data from T-P provide a new tool to investigate recruitment questions. However, this ap- proach has several limitations. First, because the geoid is not known, an estimate of mean dynamic topography (Levitus) was added to the T-P sea sur- face height anomalies. Although the accuracy of the Table 2 Mean percentage of 12-month larvae within 140 km of each bank by bank of origin, 1993-96. Bank of origin Release Destination season bank Midway Maro Necker Oahu Summer Midway 100 0 0 0 Maro 59 31 9 1 Necker 5 47 32 16 Oahu 0 28 41 31 Winter Midway 91 7 2 0 Maro 49 30 18 3 Necker 10 29 37 24 Oahu 0 19 31 50 140 Fishery Bulletin 97(1), 1999 35'N A Midway 30"N (■■■^■^M i ^ 25-N 20"N '■...■"i"a-f- % of total • >0-0.05 15"N 0.05 ■ 0.25 - 0.25-0.5 10'N O 0.5-1 0 >i -*■ :i 35"N 30'N 25'N 20'N 15"N 10'N 165"E 170"E 175"E 180" 175"W 170'W 165'W 160"W 155'W 150"W 145"W 35'N B Maro 30'N • . •, . 25'N '• N,_ ^^.; '■ .■• ?n'N ■M* ,• •-:. ► % of total • > 0 - 0.05 ; ■ ■ .'ife' 15'N " 0.05 - 0.25 o 0.25-0.5 10'N O 0.5-1 O >i '^ ::i 35"N 30'N 25'N 20'N 15'N 10'N 165'E 170'E 175'E 180' 175'W 170'W 165'W 160"W 155'W 150'W 145"W Figure 7 Simulated spatial distribution of 5000 larvae 365 days after release on 1 July 1994 at lA) Midway, iBi Maro, (C) Necker, and iDi Oahu, with an eddy diffusion rate of 500 m'/sec. The star denotes the bank from which larvae were released. Solid circles mark the other three banks. Levitus mean varied depending on the spatial and temporal distribution of the underlying data, it was likely that there were biases that may have had an impact on the accuracy of the estimates of absolute geostrophic current. A second limitation is that the spacing between data tracks, especially in mid- and low-latitude regions, may have been too broad to com- pletely resolve many important fmescale and mesos- cale physical features. Further, this tool is appropri- ate only in situations where geostrophic transport is the most significant source of larval transport, situ- ations, for example, where larvae are below the shal- low Ekman layer. In our application, this approach leads to a new hy- pothesis on the spatial dynamics of lobster larval re- cruitment in the Hawaiian Archipelago. From the simu- lations, we found that position of a spawning bank within the archipelago influences both the number of larvae spawned at a bank that will recruit to that bank as well as the recruitment that a bank will receive from other banks. Specifically, the results indicated that lar- vae are transported down the ridge from the north- Polovina et al.; Application of satellite altimetry to simulate transport dynamics of Panulirus marginatus 141 35'N C Necker 35'N 30'N 25'N 20'N 15'N 10'N % of total • > 0 - 0.05 0.05 - 0.25 0.25-0.5 . 0.5 - 1 ^ _' > 1 - 35'N 30'N 25'N 20'N 15'N 10'N 30'N 25'N 20'N 15'N 10'N 165'E 170'E 175'E 180" 175'W 170'W 165'W 160'W 155'W 150'W 145'W 35'N 30'N 25'N 20'N 15'N 10'N 165'E 170'E 175'E 180' 175'W 170'W 165'W 160'W 155'W 150'W 145'W Figure 7 (continued) west to the southeast to Necker and then southwest. Necker's persistent spiny lobster population and fish- ery appear to be the result of its location where it both receives larvae from banks to the northwest and south- east and also has a high retention of its own larvae. Only 670 km to the northwest, Maro's retention rate was only about one-half that of Necker, and although it contributed larvae to Necker, it did not receive any substantial contribution of larvae from Necker. This finding explains why recruitment at Maro remained depressed once the spawning biomass at Maro and other banks to the northwest were depleted, even though there was a substantial spawning biomass at Necker. However, it raises questions. Why were Maro and other northern banks so productive during the 1980s? Was the transport regime in the 1980s differ- ent, so that Maro and northern banks lost fewer larvae to the southeast and perhaps even received consider- able larvae from Necker? Alternatively, because the fish- ery began only in the late 1970s, was recruitment al- ways so weak at Maro and at other northern banks that the collapse at the end of the 1980s represented simply a response to overfishing an unexploited popu- lation? Although we cannot answer these questions, the results suggest that if the transport regime observed during 1993-96 persists, increased recruitment to Maro and other banks to the northwest may require an in- crease in the spawning population at Maro and other northern banks. Research on a similar species of spiny lobster, P. argus, in the Bahamas found physical trans- 142 Fishery Bulletin 97(1), 1999 35'N A July 1, 1993 35'N 30'N 25'N 20'N 15-N 10'N % of total • >0- 0.05 <■ 0.05 -0.25 0 0.25 -0.5 30"N 25'N 20"N 15'N U 0.5-1 O >1 165"E 170'E 175'E 180" 175"W 170'W 165"W 160'W 155'W 150'W 145'W 10'N 35"N 30'N 25'N 20"N 15'N B July 1, 1994 % of total • > 0 - 0.05 35"N 30'N 25-N 20"N 15'N 0.05 - 0.25 I o 0.25-0.5 IO'nI O 0.5-1 O >i 165'E 170"E 175'E 180' 175'W 170'W 165"W 160'W 155"W 150'W 145'W Figure 8 Simulated spatial distribution of 5000 lar\ au .^65 days after release at Necker on i Ai 1 July 1993, iB) 1 July 1994, and (Cll July 1995, with an eddy diffusion rate of 500 m-/sec. Solid circles denote Oahu. Maro, and Midway Islands, and the star marks Necker Island. 10"N port important to the spatial disti'ibution of larvae lLipciu.s et al., in press). However, it is important to remember that larval abundance is necessai-y but not always sufficient for good recruitment and that other factors, such as habi- tat and predators, may also be important. Further, work to improve the input parameter,? and to validate these simulation results is certainly needed. For example, the results are certainly sensitive to the lar-val depth distribution as a function of larval age. Because these simulations pioduce near-real time spatial distribu- tions, larval surveys could be designed to sample lar- vae in areas where the simulations show high and low lai'val densities in order to evaluate the model results. The model results can also be evaluated by comparing bank-specific recruitment index time series against estimated recruitment to the fi.shery three years later. Further, genetic studies based on DNA analyses may be more sensitive than the earlier electrophoi'etic stud- ies in testing the apparent lack of mixing between re- gions of the archipelago suggested by these simulations. Given the hypothesis that ti'ansport was generally from Polovina et aL: Application of satellite altimetry to simulate transport dynamics of Panulirus marginatus 143 35'N C July 1, 1995 35'N 30'N 25'N 20'N 15'N 10'N J!-5 30"N 20'N % of total ■ > 0 - 0.05 0.05 - 0.25 0 0.25-0.5 V-- ' 0.5-1 " ^ '_.' > 1 "-V 15"N 10"N 165'E 170'E 175"E 180" 175'W 170'W 165'W 160'W 155'W 150"W 145'W Figure 8 (continued) Midway to Necker, this results in a testable hypoth- esis that genetic variance should be greater at Necker than at Midway. Acknowledgments Dawn Lambeth, a NSF Research Experience for Un- dergraduates recipient, provided a great deal of as- sistance in programming and running the simula- tion model. The senior author gratefully acknowl- edges the contributions from many valuable discus- sions on satellite altimetry and physical oceanography with Professor Gary Mitchum, University of South Florida. This work was partially funded by Coopera- tive Agreement Number NA37RJ0199 from NOAA. Literature cited Gabric, A. J., and J. Parslow. 1994. Factors affecting larval dispersion in the central Great Barrier Reef In P. W. Sammarco and M. L. Herson (eds.). The bio-physics of marine larval dispersal, p. 149- 158. Coastal and Estuarine Studies ( 45 1, AGU. Washing- ton, D.C. Lipcius, R. N., and J. S. Cobb. 1994. Ecology and fishery biology of spiny lobsters. In B. F. Phillips et al. (eds.). Spiny lobster management, p. 1- 30. Fishing News Books, Oxford, England. Lipcius, R. N., W. T. Stockhausen, D. B. Eggleston, L. S. Marshall Jr., and B. Hickey. In press. Hydrodynamic decoupling of recruitment, habi- tat quality, and adult abundance in the Caribbean spiny lobster: source-sink dynamics. Mar. Freshwater Res. MacDonald, C. D. 1986. Recruitment of the puerulus of the spiny lobster, Panulirus marginatus. in Hawaii. Can. J. Fish. Aquat. Sci. 43:2111-2125. Mitchum, G. T. 1994. Comparison of the topex sea surface heights and tide gauge sea level. J. Geophys. Res. 99:24,541-25,553. 1996. On using satellite altimetric heights to provide a spa- tial context for the Hawaii Ocean time-series measure- ments. Deep-Sea Res. 43(2-31:257-280. Pearce, A. F., and B. F. Phillips. 1994. Oceanic processes, puerulus settlement and recruit- ment of the western rock lobster Panulirus cygnus. In P. W. Sammarco and M. L. Herson (eds.), The bio-physics of marine larval dispersal, p. 279-306. Coastal and Estua- rine Studies (45), AGU, Washington, D.C. Polovina, J. J., and G. T. Mitchum. 1992. Variability in spiny lobster, Panulirus marginatus, recruitment and sea level in the Northwestern Hawaiian Islands. Fish. Bull. 90:483-49. 1994. Spiny lobster recruitment and sea level: results of a 1990 forecast. Fish. Bull. 92( 1 ):203-205. Polovina, J. J., G. T. Mitchum, N. E. Graham, M. P. Craig, E. E. Demartini, and E. N. Flint. 1994. Physical and biological consequences of a climate event in the central North Pacific. Fish. Oceanogr. 3(1):15-21. Polovina, J. J., and R. B. Moffitt. 1995. Spatial and temporal distribution of the phyllosoma of the spiny lobster, Panulirus marginatus, in the North- western Hawaiian Islands. Bull, Mar. Sci. 56(2): 406-417. Seeb, L. W., J. E. Seeb, and J. J. Polovina. 1990. Genetic variation in highly exploited spiny lobster, Panulirus marginatus, populations from the Hawaiian Archipelago. Fish. Bull. 88:713-718. Shaklee, J. B., and P. B. Samollow. 1984. Genetic variation and population structure in a spiny lobster, Panulirus marginatus, in the Hawaiian archi- pelago. Fish. Bull. 82:693-702. Yu, Y., W. J. Emery, and R. R. Leben. 1995. Satellite altimeter derived geostrophic currents in the western tropical Pacific during 1992-1993 and their vali- dation with drifting buoy trajectories. J. Geophys. Res. 100(C12):25069-25085. 144 Abstract.— Monte Carlo simulation from probability distributions is often favored as a means of quantifying the uncertainty in the results of a popula- tion analysis. Observed data are com- bined with simulations from a popula- tion model by using subjective distri- butions for model parameters for which no data are available. The results from such methods can unfortunately be in- accurate unless care is taken in the combination of these simulations and the observed data. A Monte Carlo method was proposed at the 1996 meet- ing of the Scientific Committee of the International Whaling Commission for the assessment of the Bering-Chukchi- Beaufort Seas stock of bowhead whales. We show that this method is potentially inaccurate, and as such, it appears to be unsuited to the bowhead application and thus possibly to other similarly structured management problems. A proposed stock assessment method and its application to bowhead whales, Balaena mysticetus David Poole Department of Statistics P.O. Box 354322 University of Washington Seattle, Washington 98195-4322 E-mail address poole g'stat wasfiington edu Geof H. Givens Department of Statistics Colorado State University Fort Collins, Colorado 80523 Adrian E. Raftery Department of Statistics University of Washington PO Box 354322 Seattle, Washington 98195-4322 Manuscript accepted 18 March 1998. Fish. Bull. 97:144-152 ( 1999). The variability in parameter esti- mates from a population analysis is of great interest to stock assessment scientists. Monte Carlo simulation is an intuitively appealing and eas- ily applied strategy for quantifying this uncertainty. Over the past six years, various Monte Carlo assess- ment methods for the Bering- Chukchi-Beaufort stock of bowhead whales, Balaena niysticetus, have been discussed by the Scientific Committee of the International Whaling Commission (IWC). A Bayesian approach (Raftery et al., 1995) was adopted and used as the basis for the IWC assessment of the stock in 1994. The method was de- veloped after a 1991 Scientific Com- mittee (SO recommendation that methods for taking full account of uncertainty about inputs and out- puts to population dynamics mod- els be developed. An alternative maximum likelihood approach was also used for bowheads (Butter- worth and Punt, 1995; Punt and Butterworth, 1996). In contrast to the adopted method, the latter ap- proach does not allow for uncer- tainty in the values of various bio- logical parameters. Rather, it as- sumes that they are known exactly. At the 1996 SC meeting, a modi- fied maximum likelihood assess- ment to account for uncertainty in biological parameters was consid- ered (Punt and Butterworth, 1997). The assessment method was an ap- plication of a Monte Carlo approach developed by Restrepo et al. ( 1991, 1992 ). Punt and Butterworth ( 1997) cited an example of the use of the Monte Carlo approach by the Inter- national Commission for the Con- servation of Atlantic Tunas, and Restrepo et al. (1992) applied their approach to swordfish and cod fish- ery assessments. In our paper, we review the pro- posed Monte Carlo approach, both in general and in the specific bow- head application, and evaluate its performance and its compliance with established statistical prin- ciples. We show that in some cir- cumstances the method can provide suboptimal results for bowhead as- Poole et al : A proposed stock assessment method and it application to bowhead whales 145 sessment, and it may thus be unsuitable for more general fisheries management problems. In order to illustrate the potential pitfalls of the method, some simulations were performed. The paper is presented solely as a scientific appraisal of the suggested ap- proach, and the examples given are purely illustra- tive. We show how modifications of the technique can lead to improved performance, but we do not formally propose any alternative assessment methods here. The Monte Carlo approach Description Restrepo et al. ( 1992) described an approach to quan- tifying the uncertainty in the results of sequential population analyses for various fish stocks. They motivated their approach by noting that Fisheries managers recognize the dangers of accept- ing parameter estimates without consideration of the variability inherent in the estimates offish stock status and related parameters... If all sources of error are not appropriately accounted for, then estimates of the imcertainty in the assessment results may be too small. This Monte Carlo approach (hereafter MCA) proceeds as follows. Probability distributions are used to de- scribe uncertainty in the inputs to an assessment model. These distributions are constructed in two ways: 1 If observed data are available for a specific in- put, a parametric statistical model for the data is assumed, and the parameters are estimated by maximum likelihood. A parametric bootstrap is then used to obtain a sample of input values. These values are used as values of the input in the assessment model. 2 If no data relevant to the input are available, a subjective "prior" distribution, representing edu- cated guesses about the true value, is placed upon it.' A sample from this prior is used in the as- sessment model. The second case occurs in many stock assessment procedures. For instance, a prior was needed for natu- ral mortality, M, in the swordfish assessment of Restrepo et al. ( 1992). Similarly, we required a sub- jective distribution for the growth rate parameter, MSYR in the bowhead whale example in the section "Application of MCA to bowhead whale assessment." The next step in MCA is to compare simulated model outputs, such as a time trajectory of stock sizes to observed data, in order to formulate a likelihood (assuming lognormal deviations). Many input param- eter sets are drawn randomly from the specified in- put distributions (i.e. either from data-based boot- strap or subjective prior), and for each set, a condi- tional maximum likelihood estimate (MLE) is calcu- lated for quantities of interest, given the fixed input parameters and the observed data. The simulation distribution of such conditional MLEs is used to quan- tify uncertainty. The simulation is viewed as translat- ing input uncertainties into output uncertainties.'^ MCA is suggested for situations where (possibly many) nuisance parameters exist. These are typically the model input parameters for which no informa- tive data are available. The basic strategy is to esti- mate the quantities of interest (e.g. current stock size and production rate) conditional on values of the nuisance parameters ajid then to integrate over the prior for the nuisance parameters. The distribution of the conditional estimates of the parameters of in- terest is then examined for the purposes of inference. For example, if 9y is an estimator of 6, conditional on nuisance parameters y, then 9a,=leyPiY)dY (1) is an MCA estimate of 9, where p(y) is the prior for y.^ In practice, the integral in Equation 1 is not calcu- lated exactly; it is approximated by using the Monte Carlo simulation described above. In the form of an algorithm, MCA proceeds as follows: 1 Obtain the MLE 9^ from the observed data X and a likely value Yq. 2 Sample / from the prior p(7). 3 Sample pseudo-data X* from a distribution with density /"(a:; \jfiX)), where /"(x; i//) is a model for the data but not necessarily the assessment model and where i//(X) is an estimate of the parameters of this model. i//(X) may depend on the results of step 1, namely 9y and Yq, or even on standard MLEs 6 and f- An example of /"would be to as- sume X' ~ NiX, \j/), where y/ is an estimated dis- persion matrix. ' Such a distribution is effectively a Bayesian prior distribution. However, since the framework of MCA is not Bayesian, Restrepo et al. (1992) do not refer to such a distribution as a prior ' Restrepo et al. (1992) also consider uncertainty in the assess- ment model itself, but this issue is not of primary interest here. In the bowhead whale application, the assessment model is fixed by the IWC. ■' In general these parameters may be multivariate vectors. For simplicity, we restrict our focus to scalar parameters in the sub- sequent examples. 146 Fishery Bulletin 97(1), 1999 4 Find the conditional MLE 9^ given y* by using A simple point estimation example the simulated data X*. Store this value. 5 Repeat steps 2-4 many times to obtain a collec- tion of 0.^'s. MCA inference about 6 is then based on the distribu- tion of this collection of 0y,'s. Typically, the original dy^, or the mean or median of the Monte Carlo sample, is used as a point estimate, and the 0.025 and 0.975 quantiles form the bounds of a 951 confidence inter- val. The Monte Carlo approach is not a straightforward extension of a sensitivity analysis. Usually, in a sen- sitivity analysis, a small number of alternative pa- rameter values are tried, and the individual point estimates and confidence intervals obtained by us- ing each parameter set are tabled. Inference usually follows from a single analysis where a particular baseline parameter has been set; the remainder of the table is used to assess how conclusions would change under different modeling assumptions. With MCA, a potentially vast number of parameter val- ues are tried, and an overall confidence intei-val is obtained by pooling the results of each individual analysis, effectively integrating over the distribution of the nuisance parameters. We show in the rest of this paper that this integration is a source of poten- tial bias. The MCA technique is potentially highly sensitive to violations of its assumptions. If one is going to express uncertainty in the value of a parameter by means of a probability distribution, then this distri- bution should be treated as a prior in a fully Baye- sian setup. The method given in Equation 1 provides estimators that will not necessarily possess the de- sirable properties of either Bayesian or ML estima- tors. A general overview of Bayesian methods is given by Lee (1997). In a Bayesian framework, the best estimator of 0 (with respect to squared error loss) is the posterior mean E{9\ data), therefore it would be better to de- fine the estimator as 0,-2, = E[e\data) = E\E{e \y ,data)] = ^6yK[Y\data)dY, '^' where 9y - the posterior mean of 0 conditional on 7, and ;r(y|data) = the posterior distribution of y. One might regard Equation 2 as a general strategy and use it in cases where 9y is not necessarily the Bayesian estimator. In this case, however, the prop- erties of 0,2) are not clear. Schweder and Hjorf* first identified potential weak- nesses with MCA, and they described two situations where differences between the methods of Equations 1 and 2 arose. The first of these is repeated here: consider a random sample of size n from a normal distribution with mean n and variance o-, denoted X, ~ Nifi, o2) for / =: 1, . . . ,;?. Assume that there is a N(/io, xl ) prior for n, where /^q and Tq are the prior mean and variance respectively. Let cr-' be the pa- rameter for which inference is desired. ;U is a nuisance parameter and ct- is regarded as fixed. The maxi- mum likelihood estimate of d~ conditional on /j is 1 " 1=1 The estimators given by Equations 1 and 2 are then ■ 1 " and 1=1 1 " respectively, where jj^,,^, and r'^,,,, are the posterior mean and variance of the nuisance parameter ^. For this simple case, we have closed form expressions for Equations 1 and 2, and no Monte Carlo sample from the prior for /i is required. Note that in Equa- tion 2, CT~ (the conditional maximum likelihood esti- mator of (fi) is used as 9y. In addition we note that (T,2i depends on o^ because both Hp,,^, and r p^,, are functions of d^. In other words, the estimator (T,!, depends on the quantity it is trying to estimate. This occurs when 9y is not the Bayesian posterior mean of 9 conditional on y. In our examples, we simply plugged in the ordinary MLE where needed to re- move this dependency. Thus, to evaluate ct,!, here, d~j^i was used as a plug-in estimate of a- in the ex- pressions for Hp„^, and r^^,,,,. 6%^ is the usual MLE of d^ and is what would normally be used if condi- tioning on j.1 were not of interest. We know from standard theory that djf^i -> a^ as n >oo. Because r^, 6^2 1 — > CT" as n post 0 and jj post >/i, we have that ve'rge to a~ only if /Vq = P and Tq = 0. It follows that 6~it will, in general, yield accurate estimates only if ^ Schweder, T., and N. L. Hjort. 1997. Indirect and direct like- lihoods and their synthesis — with an appendix on minke whale dynamics. Paper SC/49/AS9 presented to the IWC Scientific Committee, October 1997. dm , on the other hand, will con- Poole et al.: A proposed stock assessment method and it application to bowhead whales 147 /io is close to jj and r^ is small. We show later the practical im- plications of choosing an esti- mator for which the data either do( o'l'lilordo not ( afi, ) eventu- ally dominate the prior. An alternative approach that does not involve conditioning on /J is a fully Bayesian analysis where both jj and a~ are random variables. This approach re- quires the specification of a joint prior on jj and a'^. One such prior is the indifference or "ref- erence" prior (Jeffreys, 1961) p(/j,a )-— 2-- Table 1 Simulation results for the example of Schweder and Hjort (See Footnote 4 in the | main text). 1^0 r'o MCAiff,^!, Estimators of (T^ (true value is 100) Ad hoc "Bayes": (tJ^i Full Bayes Std. MLE: alj 50 1 95.4 94.5 94.7 94.4 9 102.0 93.0 93.2 92.9 2.5 121.0 95.9 96.1 95.8 60 1 198.4 100.4 99.8 99.5 9 197.2 92.1 92.3 92.0 25 220.8 99.9 100.1 99.8 70 1 499.8 106.6 103.4 103.1 9 490.0 106.4 106.5 106.2 25 530.4 109.7 110.0 109.6 The resulting marginal poste- rior mean of cr- is the best Bayesian estimator of this parameter and is given by 1 ^(3) ■I" ( = 1 ■xV To investigate the difference between the MCA esti- mate in Equation 1 and the ad hoc "Bayes" approach in Equation 2, we performed some simple simula- tions. For each of nine combinations of ^q and Iq, a random sample of size n - 1000 was drawn from a Mp=50, 0^=100) distribution. In each case, ct,"!,, a, ^2 Cf(3) (2)' and o^ij were determined. The results are shown in Table 1. The results show very poor performance for fTm, and good, similar performances for a, i2i' ^2 C^(3i' and o^^^ . In this simple case, the analyst would presum- ably never choose MCA or the ad hoc "Bayes" method over optimal estimators, such as a^;^, and 6~ij. The key point of this example is that if conditioning on nuisance parameters is to be used, the strategy pre- sented in Equation 2 appears to be preferable to the MCA estimate in Equation 1. As mentioned earlier, the use of Equation 2 has a severe limitation of its own; we therefore do not regard it as a viable alter- native approach. There is considerable literature on the role of con- ditioning in inference. Reid (1995) has presented a review of recent developments. Confidence interval estimation The example in the section "A simple point estima- tion example" illustrates poor performance for MCA with regards to point estimation. Similar problems occur when constructing confidence intervals. Consider the following example: letX, ~ Uiy-dy, y + Oy^fori - 1 100, denote a random sample from a uniform distribution with bounds specified by the given functions of 6 and y. Let y be the nuisance parameter. The unconditional MLEs are y = ( max X, + min X, )/2 and 9 = (max X, - min X, )/(max X, + min X,). The con- ditional MLE of 0 given yis 6y- (maxX, -mmX,)/(2yi. Suppose the nuisance parameter, y, has a U(a,b) prior, 0 < a 0, we observed that MCA leads to the surprising result that the width of a quantile-based confidence interval for 9 approaches 0 as a increases, while holding the ob- served data fixed. \n other words, the width of the confi- dence interval is entirely dependent on the prior, and wider priors lead to narrower MCA confidence intervals. Application of MCA to bowhead whale assessment Punt and Butterworth ( 1997) examined the applica- bility of MCA in the assessment of the Bering- Chukchi-Beaufort stock of bowhead whales. As with the swordfish assessment of Restrepo et al. (1992), the approach involved generation of pseudodata with a parametric bootstrap. In the bowhead case, the real data consisted of abundance estimates and corre- sponding CV estimates for several years, and ob- served age-class proportions. Thus, each MCA simu- lation consisted of the following: 1 Bootstrapping of data. A series of pseudo-abun- dance estimates is bootstrapped from the observed data (Table 1 in Punt and Butterworth, 1997). Each estimate is assumed to be independent and from a lognormal distribution with mean and CV equal to the observed estimates from that year. Pseudodata for fractions of calves and matures are generated from Table 4 of IWC (1995). 2 Sampling of biological nuisance parameters from priors. Parameters such as age at maturity and natural mortality rates are generated from prior distributions from IWC (1995). 3 Conditional estimation. Conditional on the val- ues of the nuisance parameters, maximum likeli- hood estimation is used with an age-structured density dependent population dynamics model to obtain estimates of the parameters of interest: carrying capacity (K) and a productivity param- eter (MSYR). The likelihood contains contribu- tions from both the abundance and proportion data. 4 Uncertainty estimation. The variation in condi- tional MLEs is used to represent uncertainty. Note that K and MSYR are the parameters of inter- est (denoted by 9 in our previous notation), whereas the other biological parameters take the role of y. The distributions from which they were simulated are the prior distributions. The results of 1000 replications of this procedure are used to form confidence intervals. A simple population dynamics model For the purposes of illustration, we applied MCA to a simple population dynamics model (PDM). This is a non-age-structured density dependent PDM given by P,,i = P,-C, + l.5{MSYR)Pi(l-(P,/K)'), (3) where P, = the population in year t, with / = 0 cor- responding to the baseline year before commercial hunting started (here 1848); K(orPo) - the initial population size or carrying capacity; MSYR = the maximum sustainable yield rate of production as a proportion of the popu- lation aged 1+: and Cf = the number of whales killed by hunting in year t (known exactly). This model is much simpler than the BALEEN II PDM (de la Mare and Cooke^ ) used by the IWC for ^ de la Mare. W. K., and J. G. Cooke. 1993. "BALEEN II: The population model used in the Hitter-Fitter Programs". Unpub- li.'ihed manu.script available from the IWC Secretariat, The Red House, 135 Station Road, Histon, Cambridge, UK CB4 4NP. Poole et al : A proposed stock assessment method and it application to bowhead whales 149 bowhead assessment, but it nevertheless captures many of the essential features of the bowhead popu- lation. The model is viewed as having two inputs (K and MSYR) and one output (P1993); with fixed values of MSYR and the initial K, Equation 3 is applied re- cursively until P1993 is obtained. We use the term "model input" for input parameters whose true value is uncertain. Since the time series of catches C, is exactly known, we regard it as a set of constants in Equation 3, rather than as a set of model inputs. In our simplified version of the bowhead analysis, we treated MSYR as a nuisance parameter ( yin our previous notation) and K was the parameter of in- terest (the 0from before). The only parameter about which we observed data for a parametric bootstrap was P1993. MSYR was assigned a subjective prior dis- tribution. In terms of implementation, we ran the model "backwards" with P1993 and MSYR as inputs and K as output. A Newton-Raphson algorithm was used to solve for K. As a result, we effectively had P1993 and MSYR as model inputs, and K (obtained conditionally on MSYR and the data) as the model output. Implementation of MCA and a Bayesian approach To evaluate MCA, we examined its performance when the true whale stock status was known. "True" val- ues of K and MSYR were selected, and the PDM was run to obtain the "true" value of P1993. Because K (the parameter of interest here) has a "true" value, we were able to assess the accuracy of the estimates produced by the simulations. MCA was applied to the simple PDM of Equation 3 in a number of steps. The prior for the nuisance parameter MSYR was gamma(8.2, 372.7), and the likelihood for the observed total population in 1993 was A'^(Pi993,626-). These choices were based on IWC consensus (IWC, 1995) and were the same as used in previous work (Raftery et al.^ ). We assumed that we had a single observation from the likelihood for P1993. In practice such an observa- tion is usually obtained by means of a census. The observation is typically a maximum likelihood esti- mate of P1993, therefore we denoted it by P1993. An original conditional MLE was obtained by con- ditioning on a "likely" point estimate of MSYR, say MSYRq. We chose MSYRq = 0.02, the mean of the prior for MSYR. The model was then run backwards (i.e. with P1993 and MSYRq as inputs) and the likeli- hood maximized. The resulting output was the con- ditional maximum likelihood estimate K j^syrq of K because (given MSYRq) it lead to the value of P1993 that maximizes the likelihood. The MCA estimation then proceeds as follows: 1 Draw P'i993 fromM Pi993,6262). This is the para- metric bootstrap from a distribution with mean given by the observed total population in 1993. 2 Draw MSYR" from the prior for MSYR. 3 Obtain /^msvr* by running the model backwards with P'i993 and MSYR' as inputs. 4 Repeat steps 1-3 many times to form a collection of K j^fgYR* estimates. Like Punt and Butterworth (1997), we used 1000 replications. 5 Use Kjv/sy/} and the distribution of the K j^jgyn* estimates to obtain inference about K. Specifically, the distribution of the K i^syr* estimates shows how the conditional MLE of K changes as MSYR is varied according to its prior. For comparison with MCA, consider a fully Bayesian analysis which involves specifying priors for every model parameter, i.e. K, MSYR, and P1993. This in- troduces an extra complication in that the input dis- tributions and the model together induce a distribu- tion on the output. There are thus two distributions (the specified prior and the induced distribution) on the output that need to be combined or reconciled in some manner. For our "backwards" implementation of the model here, the priors for MSYR and P1993 in- duce a prior on the output K. This issue has received considerable attention at the IWC and work in the area is ongoing. A possible solution involving loga- rithmic pooling of the two distributions is discussed in Raftery et al.^ and Raftery and Poole. ^ For this example, it was useful to compare a Baye- sian approach with MCA, but avoiding the added complexity of the prior incoherence. This could be achieved if we simplified the Bayesian analysis slightly by ignoring the prior on the output K. We had a prior for MSYR and a likelihood for P1993 as we did in the MCA implementation above. In addi- tion, we now also had a M 7800,13002) prior for Pi993. This was the prior used in Raftery et al.^ and was again based on IWC consensus. The prior and likeli- hood were combined to 3deld a posterior distribution for P1993. Because we had no data on MSYR, its prior was not updated to a posterior. The only operational difference between MCA and the Bayesian method was in the generation of values for P1993: with MCA, •^ Raftery, A. E., D. Poole, and G. H. Givens. 1996. The Baye- sian synthesis assessment method; resolving the Borel Para- dox and comparing the backwards and forwards variants. Paper SC/48/AS16 presented to the IWC Scientific Committee, June 1996. Raftery, A. E., and D. Poole. 1997. Bayesian synthesis meth- odology for bowhead whales. Paper SC/49/AS5 presented to the IWC Scientific Committee, October 1997. 150 Fishery Bulletin 97(1), 1999 values were bootstrapped from a distribution whose parameters were determined by maximum likelihood; with Bayes, they were sampled from the posterior. Simulation results Simulations were performed by using three sets of "true" parameters as shown in the first three col- umns of Table 2. The values of MSYR in these three sets of simulations correspond to the 0.5, 0.975, and 0.025 quantiles (respectively) of the gamma (8.2, 372.7) prior for this parameter. In this way, we in- vestigated the performance of the method when the true MSYR is at the center and the boundaries of its prior 959^^ probability interval. In each case, a value of K was chosen such that extinction did not occur and Pi993 was positive. The first set of simulations represents a scenario where the mean of the prior for MSYR happens to coincide with the true value. If we use the prior mean as a point estimate (or "best guess") of MSYR, then our point estimate and the true value are the same. For this set, then, we would expect all assessment methods to provide accurate inference about K. The remaining two sets of simulations represent sce- narios where our prior is inaccurate, i.e. the true MSYR is either larger or smaller than the prior mean (which remains unchanged). In these cases, this in- accuracy is naturally going to cause the resulting distribution of the output K to be biased. All assess- ment approaches will be affected by this bias, par- ticularly with respect to their point estimates of K. Indeed, these parameter sets represent situations that all assessment scientists would like to avoid. The key point of interest is the extent to which a method can be insensitive to poor prior information and still provide somewhat reliable inference. The results of the simulations are shown in Table 2. For each of the three sets of true parameters, the MCA analysis was run three times by using the 0.025, 0.5, and 0.975 quantiles of the normal likelihood as the observed 1993 population, Pig^s- Then, this en- tire simulation design was replicated 500 times. The quantiles shown are MCA medians for if msyt? ■ across the 500 replicates, and the coverage rates (last two columns) show the percentage of the 500 replicates for which the estimated 95% MCA or Bayes confi- dence interval covered the truth. In the first set of simulations, where the true MSYR and the prior mean MSYRq were exactly equal, the conditional MLE /^ A/sy/?„ was very accu- rate. This is to be expected in this optimistic (if some- what unlikely) scenario. The MCA confidence inter- vals provided by K i^jgyR* covered the true K in all cases, as did the confidence intervals obtained with the Bayesian method. Also, the estimation was fairly insensitive to the accuracy of the estimate Pi^gs- A difference of 1227 whales in the estimate of P1993 (i.e. two standard deviations) resulted in if A/.sv'flo chang- ing by less than 100 whales. In the second set, the true value of MSYR was greater than the prior mean. This resulted in K having been overestimated. Here, both MCA and Bayesian results were biased by the use of the same bad prior, but the MCA coverage was worse. If follows that the subopti- mal behavior of MCA cannot be attributed solely to the choice of prior. The 95% MCA interyals provided poor coverage of the truth, worse when P1993 was accurate than when it was too low. For the Bayesian method, this difference was not as great. The coverage of the Bayesian intervals was somewhat better in two cases, particularly when P1993 was too large. In the final set, the true value of MSYR was at the low end of the prior interval, and we observed that K Table 2 Simulation results for the simple bowhead whale population dynamics model (PDM). Parameter set p ■^1993 ^ MSYRq MCA quantiles. ^msyr' Coverages MSYR K P 0.025 0.5 0.975 MCA(%) Bayesian (%) 0.02 14.700 8,733 7,506 14,640 11.230 14,360 18.870 100 100 8,733 14,700 11.260 14,400 19,160 100 100 9.960 14,780 11,330 14,450 19„540 100 100 0,04 11,300 9,896 8,669 14,700 11,2.50 14,400 19,120 68 66 9,896 14.780 11,290 14,470 19,450 55 63 • 11,123 14,880 11,580 14,560 19.840 0 23 0.01 18,500 6,971 5,744 14.570 11,270 14,300 18,430 38 54 6,971 14,620 11,260 14,340 18,710 80 88 8,198 14,670 11,270 14,380 19.000 99 99 Poole et al : A proposed stock assessment method and it application to bowhead whales 151 was underestimated by about 4000 whales for all three values of P1993. The estimates of K were inac- curate regardless of the accuracy of P 1993. Although not shown, this inaccuracy holds for the Bayesian method as well as for MCA. Both methods provided the best coverage when P1993 was too large, but the Bayesian method provided better coverage than MCA in the other two cases. Again, these results separated coverage problems attributable to inaccurate priors from additional performance degradation apparently introduced by opting for MCA analysis. As a final point, these simulations and others that we ran suggest that the MCA estimates of K are heavily dependent on the prior distribution for MSYR, to the point of being undesirably insensitive to the true values of K, P1993, and the data P1993. This behavior is more extreme than would be the case if the prior was updated to a posterior in a fully Bayesian framework. These results concur with the results in the section "A simple point estimation ex- ample," where the MCA estimates were greatly in- fluenced by the prior mean and variance. When the prior is accurate and precise, MCA may perform well, as will Bayesian techniques. Bayesian techniques seem to weight the data more heavily in relation to the prior, than does MCA. Relation between MCA and bootstrap The MCA approach described in the section "Imple- mentation of MCA and a Bayesian approach" in- cluded bootstrapping the abundance data uncondi- tionally. A standard (conditional) bootstrap would proceed by resampling the residuals (e.g. Efron and Gong, 1983, p. 43) from the model fit. The uncondi- tional approach used in the bowhead application in- troduced excess variability because bootstrapped pseudodata varied about the observed data, which in turn varied about the model fit. The general MCA approach might be viewed as an approximation to a bootstrap that is unconditional on the model fit. In such applications (e.g. Smith and Gavaris, 1993), the interpretation of the stochasticity thereby introduced must be carefully considered if it differs from the data stochasticity that causes esti- mation uncertainty. When, as is permitted with MCA, the unconditional approach simulates from a sub- jective prior rather than from data, the method is not a bootstrap because the simulation reflects stochasticity other than that introduced by data used for estimating the parameter Even in the case when sufficient data are avail- able to permit parametric bootstrap simulation of all inputs (case 1 in the section "The Monte Carlo ap- proach"), MCA does not reduce to a parametric boot- strap of the desired estimator. A parametric bootstrap expresses sampling uncertainty about a statistic RiX, F(6)), where X ~ F,hy observing the distribu- tion otRiX*, F (6)), where F is an estimate of F that depends on the data X, and X: ~ F. Variability in R is due tp X. A parametric bootstrap arises when a model F(e) = Fie) is fitted, or less desirably FiO) = G{y)m some applications of MCA. In this case, 9 or 7 should be estimated from the data, X, whose stochasticity induces sampling variability in RiX, Fie)). However, with MCA, 0 or y is estimated from different data, not the data on model output parameters, although it is the uncertainty associated with estimators of output parameters that is desired. Even in this case, the sampling distributions used are effectively data-based priors, and MCA relies on the unusual approach of integrating a conditional maximization of the likelihood over the prior. Conclusion Theoretical investigation and simulation show that the combination of Bayesian and conditional maxi- mum likelihood techniques used by MCA has the potential to yield quite variable or biased results, or both, though it can perform well in ideal circum- stances. In some situations, a fully Bayesian or clas- sical ML solution can be obtained by small modifica- tions to MCA, and the optimal properties of these more standard methods are well known. For more complex problems, Bayesian and ML solutions are sometimes more difficult to obtain than is an MCA solution. However, as our examples illustrate, MCA can result in unreliable inference even in simple situ- ations. MCA integrates a conditional maximization of the likelihood over the prior, whereas a fully Bayesian approach integrates a conditional mean. If one uses what is effectively a Bayesian prior, then it is subopti- mal to use it in a non-Bayesian inference framework. We have seen how MCA produces estimates with excessive bias. However, there may exist classes of assessment problems where, owing to some feature that is identifiable in advance, the extra bias is ac- ceptably small. MCA could be applied to such prob- lems because the excess bias would not cause MCA results to differ much from results produced by ei- ther fully Bayesian or ML methods. Our bowhead whale and simple examples clearly do not belong to such a class. Furthermore, in the general case, the extent to which MCA might err is not controllable or estimable by the analyst. Although MCA can produce good estimates in some applications, a method that can also go badly wrong is risky when one does not 152 Fishery Bulletin 97(1), 1999 know the extent of problems in any particular applica- tion. When priors are used, the usefulness of Bayesian approaches ( when obtainable) would seem to be greater than MCA, because the effect of the priors is washed out with increasing data (contrary to what happens with MCA). When priors are not required, either Baye- sian or MLE techniques might be usefially applied. Acknowledgments This research was supported by the North Slope Bor- ough, Alaska, and the National Oceanic and Atmo- spheric Administration (through the National Ma- rine Mammal Laboratory and the Alaska Eskimo Whaling Commission). The authors are grateful to Victor Restrepo and John Hoenig for comments on an earlier draft and would also like to thank Doug Butterworth, Carey Priebe, Andre Punt, Tore Schweder, Judy Zeh, and four anonymous reviewers for their helpful comments and suggestions. How- ever, the views expressed are solely the authors'. Literature cited Butterworth, D. S., and A. E. Punt. 1995. On the Bayesian approach suggested for the assess- ment of the Bering-Chukchi-Beaufort Seas stock of bow- head whales. Rep. Int. Whal. Comm. 45:303-11. Efron, B., and G. Gong. 1983. A leisurely look at the bootstrap, the jackknife, and cross-validation. The American Statistician 37:36-48. IWC (International Whaling Commission). 1995. Report of the Scientific Committee. Rep. Int. Whal. Comm. 45:53-222. Jeffreys, H. S. 1961. Theory of probability (3rd edition!. Oxford Univ. Press, Oxford, 447 p. Lee, P. M. 1997. Bayesian statistics: an introduction (2nd edition). John Wiley and Sons, New York, NY, 344 p. Punt, A. E., and D. S. Butterworth. 1996. Further remarks on the Bayesian approach for as- sessing the Bering-Chukchi-Beaufort Seas stock of bow- head whales. Rep. Int. Whal. Comm. 46:481-91. 1997. Assessments of the Bering-Chukchi-Beaufort Seas stock of bowhead whales iBalaena mysticetus) using maxi- mum likelihood and Bayesian methods. Rep. Int. Whal. Comm. 47:603-618. Raftery, A. E., G. H. Givens, and J. E. Zeh. 1995. Inference from a deterministic population dynamics model for bowhead whales (with discussion and rejoinder). -J. Am. Statist. Assoc. 90:402-430. Reid, N. 1995. The roles of conditioning in inference. Statistical Science 10:138-157. Restrepo, V. R., J. M. Hoenig, J. E. Powers, J. W. Baird, and S. C. Turner. 1992. A simple simulation approach to risk and cost analy- sis, with applications to swordfish and cod fisheries. Fish. Bull. 90:736-748. Restrepo, V. R., J. E. Powers, and S. C. Turner. 1991. Incorporating uncertainty in VPA results via simulation. Collect^ Vol. Sci.. Pap. ICCAT 35:355-61. Smith, S. J., and S. Gavaris. 1993. Evaluating the accuracy of projected catch estimates from sequential population analysis and trawl survey abun- dance estimates. In S. J. Smith et al. (eds.), Risk evalua- tion and biological reference points for fisheries manage- ment, p. 163-172. Can. Spec. Publ. Fish. Aquat. Sci. 120. 153 Abstract.— Ages of white sharks, Carcharodon carcharias. from the east coast of South Africa were estimated by counting growth rings (GRs) in verte- bral centra and relating them to length and mass. The vertebrae of 61 females (128-297 cm precaudal length (PCL)) and 53 males (142-373 cm PCL) were examined. Mass range was 42-442 kg for females (n=60) and 46-882 kg for males (n=53). X-radiography was used to enhance the visibility of the GRs in whole centra. Counts were made di- rectly from the x-radiographs by one reader and from scanned x-radiographs by two readers. Count precision for each method and reader was determined by using the average percentage error (APE ) index which ranged from 5.3 to 6.1'7c. One shark injected with oxytetracy- cline (OTC) was recaptured after 942 days at liberty. The shark was tagged at 140 cm and 46 kg and grew 69 cm and 104 kg. The OTC was visible in the vertebra and there was evidence of annual growth ring deposition. This could, however, not be confirmed with centrum analyses of the entire sample. The number of GRs counted varied in the following manner. The female and male with the lowest number of GRs, had 0 GR ( 131 cm PCL) and 1 GR (142 cm PCL), respectively. The female and male with the highest number of GRs, had 8 GRs (282 cm PCL) and 13 GRs (373 cm PCL). respectively The smallest mature male had 8 GRs (293 cm PCL); there were no mature fe- males. Von Bertalanffy parameters for the combined sexes were L^ = 544 cm PCL, k = 0.065/yr, t^ = -4.4 yr. Growth calculated from predicted lengths de- creased from 26 cm for sharks with 1 GR to 12 cm for sharks with 13 GRs. Gompertz parameters were Wg = 54 kg, G = 3.94, g = 0.094/yr Growth calcu- lated from predicted mass increased from 23 kg (1 GR) to 94 kg (13 GRs). Back-calculated lengths and mass were lower than observed values and Lee's phenomenon was evident in both back- calculated lengths and mass but not consistentlv. Age and growth determination of the white shark, Carcharodon carcharias, from the east coast of South Africa Sabine P. Wintner Geremy Cliff Natal Sharks Board Private Bag 2 Umhianga Rocks 4320, South Africa E-mail address (for S P Wintner) wintner.S'sfiark co za Manuscript accepted 30 April 1998. Fish. Bull. 153-169(1999). The white shark, Carcharodon carcharias, is found worldwide in cold and temperate coastal and shelf waters (Compagno, 1984). As a large, uncommon apex predator, it has received considerable atten- tion from scientists over the past few years. Because of its great re- sale value, compared with most other sharks, catches of white sharks have increased in various parts of the world, mainly in Aus- tralia and South Africa (Compagno, 1990, 1991). Claims of a declining population size in South Australian waters (Compagno, 1991; Bruce, 1992) influenced an initiative to implement protective legislation for this species in South Africa. In 1991 the South African government pro- hibited the catching or killing of white shark without a permit (Compagno, 1991). This measure was a preemptive protection and is to be reviewed as results of more research into the biology of this spe- cies become available (Compagno, 1991 ). Protection has now also been introduced in California, Florida, and Tasmania (Fergusson et al., in press), as well as in Queensland, New South Wales, South Australia, and Western Australia (Stevens') Legislation in South Africa has prevented this species from being targeted by commercial and recre- ational anglers. White sharks con- tinue to be caught on the east coast in the nets of the Natal Sharks Board (NSB), which operates a shark control program to protect beach users against shark attack (Cliff et al., 1988). Between 1984 and 1995 an annual average of 40 C. carcharias were caught in NSB nets, and 159^ were released alive. This activity has formed the basis of a small-scale tagging program that has provided the first esti- mates of the size of the white shark population (Cliff et al., 1996b). In- formation about the distribution, diet, movements, and catches of C carcharias in South Africa is avail- able from Bass et al. ( 1975) and Chff etal. (1989, 1996a). Little is known about age and growth of C carcharias, mainly be- cause so few are caught at any one locality, thus hampering collection of vertebral samples for ageing pur- poses. Knowledge of age at matu- rity, maximum ages, and growth rates is a prerequisite for age-based methods of stock assessment, which in turn can be used for the manage- ment of this species. The only pre- vious attempt at ageing the white shark was that of Cailliet et al. (1985) who had access to only 21 samples, mainly from California. This study attempts to provide age estimates for C. carcharias from South Africa from vertebral growth ring counts of 114 sharks. Stevens, J. D. 1997. CSIRO Marine Labo- ratories. Tasmania, Australia. Personal comm 154 Fishery Bulletin 97(1), 1999 Materials and methods Sampling Sharks were sampled in NSB nets from 1984 to 1995. Each net was 214 m long, 6 m deep, had a 50-cm stretched mesh, and was set in water 10-14 m deep, parallel to and 300-400 m from shore. For additional details of the netting operation see Cliff etal. (19881. Precaudal length (PCD was measured in a straight line from the snout tip to the precaudal notch and is used throughout this study, unless indicated other- wise. To compare our findings with those reported in the literature, the following equations were used to convert lengths: Total length (TL) = 1.251 PCL + 5.207 (n =36; range 131-307 cm PCL; 95% confidence lim- its on slope: 1.233 and 1.268; r-=0.9984) (Cliff et al., 1996a); and PCL = 0.8550 TL - 0.0955 in =58; range 96—447 cm PCL; 95'^ confidence limits on intercept: -0.130 and -0.061; r-=0.996) (Mollet and Cailliet, 1996). Lengths were converted by means of the method con- sidered most appropriate. Mass was determined by weighing each shark and subtracting the mass of gut contents where they ex- ceeded 1 kg. Maturity was assessed by the criteria of Bass et al. (1973), where males were considered mature only if their claspers were fully calcified. In females, maturity was based on the presence of dis- tinct ova in the ovary and uteri, which had expanded from a thin, tubelike condition to form loose sacs (Bass et al., 1973; Cliff et al., 1988). Vertebral samples were taken anterior to the origin of the first dorsal fin from 61 females ( 128-297 cm) and 53 males ( 142- 373 cm). Mass range was 42-442 kg for females («=60) and 46-882 kg for males («=53). Vertebrae were stored frozen (60%) or in 70% isopropyl alcohol (31%), or dried (9%). Individual centra were cleaned by removing the connective tissue from the corpus calcareum with forceps. Dried samples needed addi- tional soaking in a 5% solution of sodium hypochlo- rite for 20-40 minutes. Ring counts X-radiography was used to enhance the visibility of the growth rings. X-radiographs of whole centra were prepared on an Odel Pollux 700 generator with a Comet tube by using Agfa Ortho (Extremity) film and were processed with an Agfa Curix 160 processor We x-rayed, using mostly oblique exposure, all cen- tra with the corpus calcareum (Figs. lA and 2A) fac- ing the tube at a set distance of 100 cm. At 50 mA, exposure times ranged from 0.3 to 0.8 seconds and voltage from 28 to 34 kV. The x-radiographs were then scanned with an Agfa Arcus II scanner and Adobe photoshop software. Ulead ImagePals 2 GO! (Ulead Systems, 1992-94) and CorelDRAW! (Corel Corp., 1993) were used to enhance and work with the images. A growth ring (GR) was defined as a band pair, composed of one calcified (opaque) and one less-cal- cified (translucent) band (Fig. lA). The finer, nar- rower rings (circuli), also observed by Cailliet et al. ( 1985 ), were not used for ageing purposes. The angle change on the centrum face (Fig. 2A), a result of the difference between fast intrauterine and slower post- natal growth (Walter and Ebert, 1991), was regarded as the birth mark. Counts were made directly from the x-radiographs (XRs) by one reader (A) and from the scanned im- ages (SCs) by two readers (A and B). These three ways of viewing the vertebrae will be called meth- ods XR-A, SC-A, and SC-B, for brevity Each reader made three nonconsecutive GR counts, without knowledge of the shark's length and previous counts. Count reproducibility was determined by using the following four methods: 1 The average percentage error (APE) as described by Beamish and Fournier ( 1981) in which an up- per limit in the APE was arbitrarily set at 20% for each vertebra (samples were discarded if, af- ter a recount, they were still above this limit) and a final APE index was recalculated and an intrareader comparison (XR-A vs. SC-A) and interreader comparison (SC-A vs. SC-B) of APE values were then conducted; 2 The index of precision D ( Chang, 1982); 3 The percentage agreement among the three counts for each method; and 4 The percentage agreement in paired GR counts between the methods. Centrum analyses Dorsal centrum diameter and dorsal "birth diameter" were measured in a transverse plane along a straight line through the focus of each vertebra ( Fig. 2B ). The dorsal "birth diameter" was marked on the x-radio- graphs and screen images. Distance from the focus to the outer edge of each GR (Fig. 2B) was measured on the scanned images by using CorelDRAW! Wintner and Cliff: Age and growtfi determination of Carcharodon carcharias 155 3.32 cm Dorsal diameter B ^m 3.70 cm ^r^ OB TB OB TB OB OTC location under UV light TB Birth mark ^ GR Zl OR GR R.B84044 9/8/84 Female 191 cm 115kg Figure 1 Inverted scanned images of x-radiographs of two white sharks. OB = opaque band, TB = translucent band. GR = growth ring. Tagged: Recapture: BT433 AMA97014 30/10/94 27/5/97 Male 140 cm 209 cm 46 kg 150 kg 13.0 ml OTC Dorsal Intermedialia Caudal Angle change (=birth mart) Ventral Dorsal "birth diameter" Corpus calcareum B Dorsal Ventral Figure 2 Schematic drawings of a vertebra illustrating terms used in the text. Distances from focus to outer edge of each growth ring Dorsal centrum diameter The relationships between centrum diameter and both shark length and mass were examined by com- paring regression lines of both sexes with the proce- dure of Zar (1974), which compares the slopes and elevations with Student's f-tests. Significance levels given in the text are the results of these tests. Outli- ers were determined by using Statgraphics (STSC, 1991) and eliminated. The Dahl-Lea method of back-calculation (Car- lander, 1969) was used, in which PCL, = CD, (PCLJCD^), where PCL, = length at GR t\ CD, - centrum diameter at GR t\ PCL^ - length at capture; and CD^ = centrum diameter at capture. The Monastyrsky method of back-calculation (Bagenal and Tesch, 1978, cited by Francis, 1990) was used to estimate mass at age, in which M, = iCD/CD/ M^, where M, = mass at GR t; 156 Fishery Bulletin 97(1), 1999 M = mass at capture; and b - the constant derived from the multi- pUcative regression of M on CD. The constant b was derived by using the "body-pro- portional-hypothesis," where "if a fish at time of cap- ture was 10 per cent smaller than the average fish with the same size of scale, the fish would be 10 per cent smaller than the expected length for the size of that scale throughout life" (Whitney and Carlander, 1956, cited by Francis, 1990). Confirmation of the annual periodicity of GRs (Cailliet et al., 1983a, 1983b) was attempted with two methods of centrum analyses. First, the last de- posited band was classified as translucent or opaque and related to the month of capture (Kusher et al., 1992). The observed and expected ratios of translu- cent to opaque last bands were then compared. Sec- ond, the marginal increment ratio (MIR) (Hayashi, 1976; Skomal, 1990) was calculated with the follow- ing equation: MIR = (VR-R)/{R-R n-l where VR = vertebral radius; radius to the last complete GR; and radius to the previously completed GR. Mean MIR, with range and standard error, was then plotted against month. In addition, 16 white sharks (between 1993 and 1995) were injected with oxytetracycline (OTC) so- lution (Engemycin, Intervet International B.V.), at a dose of 25 mg per kg body mass as recommended by Holden and Vince (1973) and McFarlane and Beamish ( 1987). Mass was estimated from the mass- length equation of Cliff et al. (1989). The OTC was administered in several places in the muscle around the first dorsal fin with a 15G x 1.5" disposable needle and 20-mL plastic syringe. Each shark was tagged with an orange identification tag (Hallprint PDA large plastic dart tag), labelled "tetracycline" and given a unique "BT" number. Age and growth The program PC-YIELD II (Punt and Hughes, 1989) was used to determine which of 10 different growth models provided the best fit to the data sets obtained by the three methods. Whe.re appropriate, von Bertalanffy growth parameters (VBGP) were com- puted. The equation (von Bertalanffy, 1938) is L, ^L^d-e-*"-'"'), Table 1 Comparison of average percer tage error ( APE ) indices and inde.x of precision {D) for each method. Values are given before and after elimination of vertebrae with an APE > 0.2. Preliminary Final APE index APE index Methods ('7i) D{9c} n (7f) DCX) n SC-A 8,9 6.9 114 5.3 4.1 112 SC-B 12.9 9.8 114 5.4 4.1 108 XR-A 9.0 7.0 114 6.1 .3.9 110 where L^ = length at GR t; maximum theoretical length; the rate at which L^ is reached; and the theoretical number of GR at length zero. k" = t„ = To determine whether GR deposition was related to an increase in mass rather than to time of year, Gom- pertz growth parameters were also calculated. The Gompertz equation (Silliman, 1967; Ricker, 1975) is W,= u'oe«'i- where w^ = mass at GR t; G - initial exponential growth; and g = exponential rate of decline. Both growth equations were fitted by using the nonlin- ear regression procedure of STATGRAPHICS, which uses Marquardt's algorithm (Draper and Smith, 1981). Because this procedure is highly dependent upon ini- tial estimates for the parameters, a wide range of ini- tial parameters was used to prevent the models con- verging to a local minimum, i.e. converging to an un- wanted stationary point of the sum of squares, rather than to a global minimum (Draper and Smith, 1981). Results Of the 1 14 processed vertebrae, between two and six, depending on the method, had an APE of over 209^ after a recount and were discarded (Table 1). The final APE indices and D values showed little differ- ences amongst methods, indicating similar reproduc- ibility (Table 1). The percentage agreement among the three counts was high, e.g. with method SC-A, for 58. 0'^ of the sample the three counts were the same (Table 2). In all three methods the majority of the readings was the same or differed only by one GR. For this reason, a mean of the three counts was Wintner and Cliff; Age and growth] determination of Carcharodon carcharias 157 7 -1 6 - + $^ f ^^ 5 - meter (cm) 1 3- intercept -0 284 (SE0 082) g slope 0 0188 (SE 0 0004) c n 114 ^ 2- 1 - •r ■ r 0 96 0 - 1 1 1 1 1 1 ) 50 100 150 200 250 300 1 1 350 400 Precaudal length (cm) Figure 3 Relationship between centrum diameter and length for combined sexes of the white shark , C. carcharias. Squares indicate shark BT433 at tagging and recapture. Table 2 Percentage agreement among the three counts for each method. Numbers in parentheses indicate sample sizes. Difference in GR counts Methods 0 GR 1 GR 2 GR 3 GR 4 GR n SC-A 58.0 (65) SC-B 61.1(66) XR-A .50.0(55) 33.0(37) 7.1(8) 0.9(1) 0.9(1) 112 37.0(40) 0.9(1) 0.9(1) — 108 36.4(40) 11.8(13) 1.8(2) — 110 Table 3 Percentage agreement in paired growth ring (GR) counts between methods. Numbers in parentheses indicate sample Difference between counts Methods OGR 1 GR 2 GR 3 GR SC-A- SC-A- SC-B XR-A 42.5(45) 59.6(65) 41.5(44) 34.9(38) 15.1(16) 4.6(5) 0.9 0.9 106 109 taken as a GR estimate in all three methods and used for all further calculations. The two readers agreed on the mean GR estimate in 42.59c of the vertebrae; in 59. 6% of the vertebrae there was no difference between GR estimates obtained by the same reader using the two methods (Table 3). Centrum analyses Prebirth marks were found in all vertebrae with method XR-A and in 33% of the vertebrae with method SC-A (Table 4). Using all three methods, we Table 4 Percentage of vertebrae where prebirth marks were present and nature of the first band after the angle change. Method SC-A SC-B XR-A Prebirth marks First band after angle change opaque translucent % 33 100 114 114 94 96 6 4 12 62 97 81 158 Fishery Bulletin 97(1), 1999 Table 5 Observed and back-calculated prec audal length (PCL) at number of growth ring for the white shark. C. carch arias. G rowth ring estimates were obtained with method SC-A. No. of Observed PCL (cm) Back-calculated PCL (cm) growth rings Min Max Mean SD n Min Max Mean SD n 0 131 131 131 1 85 118 100 7 114 1 128 178 156 16 12 96 182 133 17 112 2 161 228 192 19 18 121 221 165 22 99 3 172 236 204 17 25 150 241 191 24 77 4 197 265 225 16 26 169 270 212 25 46 5 215 291 246 22 15 184 261 222 21 22 6 238 297 276 21 7 212 271 240 18 13 7 263 307 287 19 4 233 294 264 19 10 8 282 293 288 8 2 259 301 279 18 5 9 — — — — — 278 308 289 16 3 10 317 317 317 — 1 300 335 317 24 2 11 — — — — — 325 325 325 — 1 12 — — — — — 338 338 338 — 1 13 373 373 373 — 1 350 350 350 — 1 14 — — — — — 369 369 369 — 1 found that the majority of the vertebrae had an opaque band as the first band after the angle change (Table 4). Only samples where the nature of the first band was the same in all three counts were consid- ered for our analysis. A statistically significant linear relation was found between centrum diameter and PCL (Fig. 3). There was no significant difference between the sexes in the slopes (P>0.5), but a significant difference be- tween the elevations (P<0.001). Because the inter- cepts differed only by 0.02, no visual difference in the predicted values of females and males was per- ceptible, and the common intercept was used to plot a regression for the combined sexes. The intercept was close to zero; therefore no correction, such as the Fraser-Lee method (Carlander, 1969; Branstetter, 1987), was used. One of the 16 white sharks injected with OTC (BT433) was recaptured after the completion of the study. The shark was tagged during a fishing com- petition on 30 October 1994; it measured 140 cm and weighed 46 kg. On recapture 942 days later, on 28 May 1997, it measured 209 cm ( 150 kg). The squares in Figure 3 indicate the length and centrum diameter of this shark at tagging, i.e. at the OTC marker and at recapture. Because of the similarity of the above regressions for females and males, back-calculations were per- formed on combined sex data. Mean back-calculated lengths were lower than observed values (Table 5). Lee's phenomenon, a tendency for back-calculated lengths of older fish in the earlier years of life to be systematically lower than those of younger fish at the same age (Carlander, 1969; Smith, 1983), was evident, but not consistent. In sharks with 7 and 8 GRs, for example, the back-calculated values for each GR class were higher than those for sharks with 6 GRs. A multiplicative relationship was found between centrum diameter and mass and no significant dif- ference was found between the sexes in the slopes (P>0.5), but a significant difference between the el- evations (P<0.001) (Fig. 4). Again, because of the similarity of the regressions for females and males, back-calculations were performed on combined sex data. Back-calculated mass was lower than observed values (Table 6). Lee's phenomenon was evident, but not consistent, and showed a similar trend to that of back-calculated lengths. Because of the use of mean GRs, the number of observed lengths (or masses) per GR class, plus the number of back-calculated lengths (or masses) in the next GR class, did not always add up to the total number of back-calculated lengths (or masses) of the previous GR class (Tables 5 and 6). The observed ratio of translucent to opaque last bands differed significantly from the expected ratio (X^ test, P<0.001, /!=33), irrespective of whether opaque band deposition was assumed to occur in sum- mer or in winter. For this analysis only vertebrae were used where the nature of the last band was the same in all nine counts (i.e. in all three counts of all three methods). The exclusion of borderline cases, i.e. only summer months (22 Dec-20 Mar) and win- ter months (21 Jun-22 Sepj, did not alter the result Wintner and Cliff: Age and growthi determination of Carcharodon carchanas 159 7- 4 - I - - Female Male y^ax"; slope=b. elevation= ln(a) | common slope 0 366 (SE 0 009) Females Males: intercept ■0 539 (SE 0 054) -0 529 (SE 0 065) f 0 95 094 n 60 51 100 200 300 400 500 Mass (kg) 600 700 800 900 Figure 4 Relationship between centrum diameter and mass for female (xl and male (o) white sharks, C. carchanas. Squares indicate shark BT433 at tagging and recapture. Table 6 Observed and back-ca culated mass at number of growth ring for the white shark, C. carcharias Growth ring estimates were obtained mth method SC-A. No. of Observed mass (kg) Back-calculated mass (kg) growth rings Min Max Mean SD n Min Max Mean SD n 0 42 42 42 1 14 36 22 4 112 1 46 97 69 19 11 18 120 48 17 110 2 68 199 120 37 18 33 198 85 31 98 3 81 218 139 35 25 55 256 124 44 76 4 98 250 176 42 26 73 352 164 55 46 5 112 368 234 68 15 90 280 188 48 22 6 218 442 314 84 7 186 313 239 42 13 7 272 442 386 77 4 238 416 310 55 10 8 371 437 404 47 2 351 406 380 23 5 9 — — — — — 420 504 452 45 3 10 544 544 544 — 1 506 625 565 84 2 11 — — — — — 618 618 618 — 1 12 — — — — — 685 685 685 — 1 13 882 882 882 — 1 751 751 751 — 1 14 — — — — — 859 859 859 — 1 iX^ test, P<0.001, n=22). With method SC-A, and again only with vertebrae where all three counts had the same last band, the observed ratio of translu- cent to opaque last bands differed significantly from the expected ratio (x^ test, P<0.001, n=75), again ir- respective of whether opaque band deposition was 160 Fishery Bulletin 97(1), 1999 + -•-Mean - Min 2 - + + Max * • . • 1.5 - * 'y^ 0.5 • ~ — 0 • 1 1 1 . 1 ■ . 1 ■ < > < 0 1 2 3 4 5 6 7 8 9 10 11 12 13 Month Figure 5 Marginal increment ratio (MIR) by month for the white shark, C. carcharias. Numbers indi- cate sample size, vertical bars indicate the standard error of the mean, and +/- indicates the range. Measurements were obtained by using method SC-A. 1 = January. 12 = December assumed to occur in summer or in winter. The exclu- sion of borderline cases, however, resulted in a sig- nificant difference between observed and the ex- pected ratios (x' test, P<0.001, n-39), assuming opaque band deposition in summer, but no signifi- cant difference (X" test, P>0.05, n=39), assuming a translucent band deposition in summer. The vertebra of shark BT433 was difficult to read (Fig. IB), when compared with other sharks of simi- lar length and mass (Fig. lA). For this reason, we also examined the vertebra with transmitted light (Wintner and Cliff 1996), and our interpretation of GRs is based mainly on this method. The OTC marker was visible in the opaque band (injected 30 October 1994). The deposition of OTC when injected intramuscularly occurs after a couple of weeks (Holden and Vince, 1973) or 21-35 days (Brown and Gruber, 1988). The opaque band would therefore have been deposited in summer (November or December). The last band in the vertebra was an opaque band. Because the shark was recaptured on 27 May 1997 (the beginning of winter), the translucent band might have been in the process of being formed. MIR analysis of the entire sample (Fig. 5), how- ever, did not show a distinct time of GR formation because mean and minimum ratios did not get close to zero. The results of a Kruskal-Wallis analysis in- dicated that there is no relation between MIR and month (P=0.39). In view of the results from the above recaptured shark and if the peak in July was re- garded as a single peak, we assumed that one GR was formed annually, combined with the minimum MIR trend. With the latter, GR formation may occur during December or January. Because of the ambi- guity of the results of the centrum analyses, how- ever, annual periodicity of GR could not be confirmed. Age and growth The von Bertalanffy growth function was the most appropriate model for the data sets obtained by both methods SC-A and XR-A. The length-GR data set obtained by method SC-B was not investigated fur- ther because two growth models. Putter no. 2 and Gompertz (Punt and Hughes, 1989), provided a bet- ter fit than did the von Bertalanffy growth function. In addition, the L^ values obtained by the two mod- els were too low to be realistic according to observed sizes of white sharks. Sexes could not be compared because there were no females larger than 300 cm; consequently no meaningful von Bertalanffy growth function could be fitted. Although the VBGP for the two methods SC-A and XR-A differed, there was little difference in the calculated GR over the range of lengths sampled (Fig. 6). Method SC-A had lower relative standard errors and a more realistic L than Wintner and Cliff: Age and growth determination of Carcharodon carcharias 161 ■a •o 400 350 300 - 250 - 200 100- 50- -SC-A XR-A Parameters SC-A Parameters XR-A Lrf(cm) 544 (SE 121) 686 TL 665 (SE 238) 837 TL k (yr') 0 065 (SE 0 026) 0 043 (SE 0 024) t<,(yr):-4,4(SE0 8) -5 3(SE1 1) n: 112 110 r^: 0.83 0 82 5 6 7 8 Number of growth rings 10 12 13 14 Figure 6 Von Bertalanffy growth curves and parameters for the white shark, C. carcharias, sexes combined. Growth ring estimates were obtained by using methods SC-A and XR-A. Observed values { + ) are those for the method SC-A. Squares indicate shark BT433 at tagging and recapture. method XR-A. For these rea- sons the results of the former method were used for back- calculations, MIR analysis, fit- ting of the Gompertz growth curve, growth calculations, and GR estimates. Table 7 shows the number of GRs counted for various animals. The largest female, an adolescent, (297 cm) had 6 GRs; there were no mature females. Of the three mature males, the smallest had 8 GRs (293 cm), the other two had 10 and 13 GRs (317 and 373 cm), respectively. Size at birth ranged be- tween 100 cm (back-calculated value) and 135 cm (predicted value). Shark BT433 was found to have one GR at tagging ( 140 cm) and three GRs at recap- ture (209 cm), after 2.6 yr at liberty (Fig. IB; Fig. 6). A Gompertz growth curve was fitted to the mass- GR data ( Fig. 7 ). The female and male with the high- est number of GRs, had 8 GRs (437 kg) and 13 GRs (882 kg), respectively (Table 7). Mass at birth ranged Table 7 Growth ring (GR) counts for animals of certain lengths (PCL, cm) or certain mass (kg). Male Female No of GRs PCL mass No of GRs PCL mass Lowest number of GRs 1 142 56 0 131 42 Highest number of GRs 13 373 882 8 282 437 Smallest animal 1 142 46 1 128 42 Largest animal 13 373 882 6 297 442 Smallest mature animal 8 293 371 — — — between 22 kg (back-calculated value) and 54 kg (pre- dicted value). Mean growth rates calculated from observed lengths were higher overall than those calculated from predicted lengths (Fig. 8). Mean growth for pre- dicted lengths was 26 cm for the period 0-1 GR, de- creasing to 20 cm (4-5 GRs), 16 cm (7-8 GRs), and 12 cm ( 12-13 GRs). Mean observed growth in mass was also higher than that of predicted values (Fig. 8). Mean growth for predicted mass was 23 kg for 0- 162 Fishery Bulletin 97(1), 1999 QOA yuu - 800 - 700 ■ 600 ■ y^ + ■S, 500 ■ S 400- + + y^ + y^ + 300 - + + ^^ + ^/^ Parameters: i i^ ' w„ (Kg): 54 (SE 7) G: 3 94 (SE 0 29) 200 - + i ^^* + g(yr'): 0.094 (SE 0 018) n: 111 c^: 0.84 100 - : — m U T 1 1 1 1 1 r- 1 1 1 1 1 1 1 0 1 2 3 4 5 6 7 8 9 10 11 12 13 Number of growth rings Figure 7 Gompertz growth curve for the white shark, C. carcharias. sexes combined. Growth ring estimates were obtained by using method SC-A. Squares indicate shark BT433 at tagging and recapture. 1 GR, increasing to 51 kg (4-5 GRs), 78 kg (7-8 GRs) and 94 kg (12-13 GRs). Shark BT433 grew 69 cm and 104 kg in 2.6 years (Fig. IB) which represents a mean annual growth of 27 cm and 40 kg. Discussion Annual GR periodicity has been partly verified for several species (Cailliet, 1990). For some species of the family Lamnidae, biannual GR periodicity has been reported. Parker and Stott ( 1965) used the mean length of 17 Cetorhinus maximus sampled in winter and 15 sampled in summer and treating them as age classes, derived a tentative growth curve. They stated that their growth curve derived from ring counts of live vertebrae was similar to the first one when a depo- sition of two GRs per year was assumed. They were, however, careful to note that there was "nothing to in- dicate beyond doubt that the addition of two rings is the direct result of the passage, of an annual seasonal cycle" and said that several findings did not "harmonise with the idea of annual increase of two rings." Pratt and Casey ( 1983) determined age and growth of Isurus oxyrinchus from the Atlantic by length- month analysis, length-frequency analysis, tag-re- capture information, and vertebral ring counts. The results of the first analysis were used to interpret the accuracy of the other methods. Their growth curve for /. oxyrinchus, based on back-calculated sizes, agreed closely with those obtained by the other methods if biannual ring deposition was assumed. The vertebrae from four noninjected recaptured /. oxyrinchus, however, gave inconclusive results be- cause two supported annual and the other two bian- nual GR deposition. Pratt and Casey stated that "ver- tebral rings may thus yield an approximation of age, accurate in the smaller sizes where estimates have been correlated with other methodologies. Adults may not lay down yearly, and it is possible that we have underestimated their age, but we have no data to support this possibility." Cailliet et al. ( 1983a) assumed annual GR deposi- tion in /. oxyrinchus from California waters. Their growth rate estimates, based on tag-recapture analy- sis, were therefore half of those of Pratt and Casey (1983) and had a much smaller variation in the esti- mate. Cailliet et al. ( 1983a) stated that although "this discrepancy could be related to differences in habi- tat or environmental conditions or differences in Wintner and Cliff: Age and growtfi determination of Carcharodon carcharias 163 ■S s o 100 90 80 - 70 I 1 Weight ■^■Length Weight Length 5 6 7 8 Number of growth rings 10 12 13 Figure 8 Observed ( bar ) and predicted (line ) growth in mass and length per growth ring for the white shark, C. carcharias. Growth ring estimates were obtained by using method SC-A. sample size or ageing methodology" it was interest- ing to note "that the growth rate reported by Pratt and Casey (1983) based on their back-calculation from counts of bands on centra, would be similar to ours if each pair of bands from their fish were inter- preted as an annual event." Cailliet et al. ( 1985) assumed annual GR periodic- ity in their age and growth study of C. carcharias. As with the studies mentioned above, MIR analysis could not be included to validate the temporal peri- odicity of the GRs. The only lamnoid study to have done so, apart from this one, was that of Branstetter and Musick (1994) on Carcharias taurus, which sug- gested a semiannual periodicity of band and ring for- mation. However, samples from three winter months were lacking. They also used a "odd-even ring count analysis" to verify this suggestion. The results of the MIR analysis (Fig. 5) were in- conclusive and did not confirm the results fi-om shark BT433. Considerable time was spent on this analy- sis and great care was taken to discern the last de- posited band in order not to overlook a recently formed band. Several vertebrae were remeasured, resulting only in removing the peaks of the curve and not in a reduction of the minimum MIR. Exami- nation of the relative frequency of vertebrae with large MIR to those with low MIR plotted against month (Batista and Silva, 1995) did not shed light on this issue. Annual or biannual GR periodicity for C. carcharias could not be confirmed in this study by two centrum edge analyses. There seems to be some unexplained variation in GR deposition among lamnoids, which is further com- pounded by our study Pratt and Casey ( 1983) stated that in /. oxyrinchus instead of the traditional an- nual ring deposition "a more likely cause for ring for- mation would be times of stress or deprivation such as migration and mating." Similarly, Branstetter and Musick (1994) suggested that in C. taurus the for- mation of semiannual bands may reflect their north- south seasonal migration pattern, which is prompted in part by changing light and temperature patterns. There is currently not enough evidence to prove a similar migration pattern for C. carcharias in South African waters (Cliff et al., 1996a, 1996b). Traditionally, ages of sharks have been related to length (Cailliet et al., 1983b; Cailliet et al., 1986). Natanson and Cailliet (1990) found that band depo- sition in Squatina californica was not annual but related to somatic growth. We therefore decided to fit a Gompertz growth curve because this curve usu- ally describes the relationship between mass and age 164 Fishery Bulletin 97(1), 1999 well (Ricker, 1975). The curve did not show the typi- cal asymmetrical sigmoid curve but seemed to ap- proach an upper asymptote (Gulland, 1983). It was the method with the lowest relative standard errors for all parameters. Although there is also consider- able variation in mass at GR, we felt that the Gompertz growth curve has merit, especially because there is a large change in mass associated with a small increase in length among large sharks. The results of the MIR analysis could be interpreted such that GR formation is not related to time of year but to mass increase (some sharks taking a longer time than others to gain the same amount of mass). Typically, opaque band deposition is associated with summer growth (Cailliet et al., 1983b, 1986; Kusher et al., 1992), and the nature of the last de- posited band can be related to the month of capture to verify this. Using method SC-A and only summer and winter months, we found that the observed ra- tio of translucent to opaque last bands did not differ significantly from the expected ratio, assuming a translucent band deposition in summer. When we used nine ring counts, however, the analysis did not show any relation between the nature of the last band and season. The analysis could be considered statis- tically weak owing to the low sample size but was included to emphasize the accuracy of the last band identification. In this study the band immediately after the angle change was opaque in most verte- brae, which is in keeping with Francis (1996) who reported time of parturition of white sharks as spring or summer. In addition, if our interpretation of ver- tebral bands in shark BT433 is correct, the opaque band would be formed in summer Because of the inconclusive results of the centrum analyses of the entire sample, however, more recaptures of sharks injected with OTC are needed to confirm this theory. The relation between centrum diameter and shark length was linear, as was found in C. carcharias by Cailliet et al. ( 1985) and in several other shark species (Cailliet et al., 1983b; Schwartz, 1983; Branstetter, 1987). For another species of the family Lamnidae, /. oxyrinchus, Pratt and Casey (1983) found a slightly curvilinear relationship for both females and males. The relation between centrum diameter and mass was multiplicative, which explains the slightly big- ger difference in centrum diameter between the sexes than in the relation between centrum diameter and length. The differences between the sexes in both relationships, however, are slight and are probably not of biological significance. Back-calculated mass and length values were lower than observed values. The differences between mean observed length (or mass) and mean back-calculated length (or mass) at each GR would decrease substan- tially if the observed GR 0 were treated as GR 1. This could be an indication that the angle change of the corpus calcareum is not formed at birth but in the first summer growth (Brown and Gruber, 1988; Wintner and Cliff 1996). X-radiography to enhance the visibility of GRs in elasmobranch vertebrae has been used successfully on several species (Cailliet et al., 1983a, 1983b; Yudin and Cailliet, 1990; Ferreira and Vooren, 1991). This technique was also used in the only other ageing study of C. carcharias by Cailliet et al. (1985). They counted GRs directly from x-radiographs and used silver nitrate staining to corroborate counts of larger vertebrae that proved more difficult to read. In our study, scanned images allowed for easy and rapid counts of rings and measurements for back-calcula- tions. In addition, scanned images were easier to in- terpret because they showed less detail of the narrow circuli and prebirth marks thcin did the x-radiographs. Prebirth marks in placental species are normally attributed to the time of placenta formation and at- tachment (Casey et al., 1985; Branstetter, 1987; Branstetter and Stiles, 1987). In our study, prebirth marks were found in C. carcharias vertebrae. No comments on prebirth marks in C. carcharias, I. oxyrinchus, or Alopias vulpiniis were made by Cailliet et al. (1985), Pratt and Casey (1983), and Cailliet et al. ( 1983a), respectively Branstetter et al. ( 1987 ) did not find prebirth marks in Galeocerdo cuvier, another aplacental species. Branstetter and Musick (1994), however, found prebirth marks in their Carcharias taiirus specimens and related the first consistent prebirth ring to the size of the embryo when diges- tion of the large quantities of eggs begins. We did not relate prebirth marks to embryonic length using back- calculations because they represent growth in utero. The APE indices for the three methods (5.3-6.1%) were considered acceptable. They were lower than those of the four methods used by Wintner and Cliff ( 1996) for Carcharhinus limbatus (8.1-13.0%, «=80- 87) and those of Cailliet et al. (1990) who used "bow tie" sections of Miistelus manazo (6.9-12.7%, n=28- 30). D-values (3.9-4.1%) were also considered to be acceptable because they were similar to those of Natanson and Kohler (1996) for C. obscurus (3,3%, 7i=42) and lower than those of Cailliet et al. (1990) (6.8-12.7%, n=27-30). Age and growth estimates Only one shark injected with OTC was recaptured (BT433). Although it was at liberty for an adequate time period, an interpretation of the bands on the x- radiograph was difficult, and therefore the results were based mainly on viewing the vertebra with Wintner and Cliff: Age and growtfi determination of Carcharodon carcharias 165 transmitted light. The num- ber of GRs counted, however, were the same with these two methods and for both authors. While at liberty, the shark's mean annual growth increment was 28 cm and 42 kg. Francis (1996) reported size at birth between 92 and 116 cm and weight at birth between 12 and 32 kg and given that this shark was 140 cm and 46 kg at tagging, it is reasonable to assume that its age was in the region of one year. This estimation was confirmed by the OTC marker being visible at the edge of the first GR (Fig. IB). Because we counted three GRs at recapture, the mean predicted growth/GR in the period from 1 to 3 GRs was 24 cm/GR and 33 kg/ GR (Fig. 8). Assuming that one GR is deposited per year, the total growth of shark BT433 would be 62 cm and 86 kg in the period of 2.6 years. If the GR deposition is biannual, which is not impossible given the difficulty in interpreting this vertebra (Fig. IB), the mean predicted growth would be 18 cm/GR and 32 kg/GR; this would amount to a total growth in a period of 2.6 yr (5.2 GR) of 94 cm and 166 kg. Be- cause BT433 grew 69 cm and 104 kg, the first of these two predictions fits the observed growth better than the second. Further evidence to support the hypothesis of an- nual GR deposition can be found in Figures 6 and 7. If the number of GRs represents years, the expected size and mass of shark BT433 at tagging and recap- ture after 2.6 years is in accordance with the observed values of our sample (Fig. 6). Assuming biannual GR deposition, we believe the shark's size and mass at tagging is still in accordance with the observed val- ues (Fig. 7); however, it would have been in the re- gion of 288 cm and 375 kg at recapture. Because evi- dence of annual GR deposition in C. carcharias is based on the recapture of a single shark injected with OTC and because our centrum analyses neither confirmed nor contradicted annual ring deposition, the discussions below are based on number of GRs rather than years. Our VBGPs, in the absence of very large sharks, were L^ = 544 cm (SE 121), k = 0.065/yr (SE 0.026), and tg = -4.4 yr (SE 0.8). We fitted a von Bertalanffy growth curve to the data points presented by Cailliet et al. (1985) in order to compare standard errors (Table 8). The results were very similar, but given the larger sample size in our study, it would appear Table 8 Comparison of von Bertalanffy growth parameters. The bold values were obtained by | refitting the data. Our study Cailliet et.aK 1985) (SC-A) Their results Refitted excluding 3 large animals Range PCL 128-373 110-^34 110-394 Range TL 165-472 129-508 129-461 VBGP L 544 (686 TL) 653 (764 TL) 569 (666 TL) SE 121 134 (157 TL) 110(129TL) k 0.065 0.058 0.072 SE 0.026 0.023 0.028 'o -AA -3.5 -3.3 SE 0.8 0.72 0.72 n 112 21 18 r2 0.83 0.98 0.97 that there is greater variation in length at number of GRs in C. carcharias from South Africa. Our largest shark (373 cm) measured 452 cm TL. The absence of larger sharks in our study undoubt- edly accounts for the lower L^ of 544 cm (686 cm TL), as opposed to that of Cailliet et al. (1985) of 654 cm (764 cm TL). This finding was confirmed when the three markedly larger sharks (494-508 cm TL) of their study were omitted and the recalculated L^ was 569 cm (666 cm TL) (Table 8). The conversion of the original data of Cailliet et al. (1985) from TL to PCL had no effect on the VBGP and their standard er- rors. One should keep in mind, however, that com- parisons of L^ are somewhat hampered by the fact that TL length is measured in two different ways (Mollet et al, 1996) by various authors, although the difference in C. carcharias is not as pronounced as, e.g., in members of the family Carcharhinidae. The maximum size attained by the white shark is the subject of much interest and controversy and, more importantly, uncertainty (Ellis and McCosker, 1991). Randall (1973, 1987) refuted the lengths of 1113, 900, and 640 cm TL attributed to C. carcharias. According to him, the largest reliable measured white shark is 513 cm (600 cm TL). Mollet et al. (1996) calculated the size of two large white sharks, using three morphometric measurements, at 453-701 cm (530-820 cm TL) and 393-598 cm (460-700 cm TL), respectively. These results were consistent with the estimated TL of >700 and 700 cm, respectively. They concluded that the most solid TL estimates for these two sharks were those original estimates. Our L^ is larger than the shark from Randall and is smaller than the data of Mollet et al. ( 1996) and Cailliet et al. (1985). 166 Fishery Bulletin 97(1), 1999 Our growth coefficient is similar to that found by Cailliet et al. ( 1985), and both are an order of magni- tude lower than that of 7. oxyrinchus at 0.203-0.266 (Pratt and Casey, 1983) and that oiLamna nasus at 0.116 (Aasen, 1963). In the first year of growth, C. carcharias grows 19-26% of its size at birth which is less than that of other lamnoids (Branstetter, 1990). Our findings place C. carcharias in Branstetter's (1990) group of sharks with slow growth (/?<0.10; growth in the first year<30% of the birth size). In our study, back-calculated size at birth was 100 cm and calculated value was 135 cm. The smallest accurately measured free-swimming C. carcharias in southern Africa was 108 cm (140 cm TL) (Smith, 1951) and the sizes of the smallest white sharks caught at the NSB are 128-145 cm. "Umbilical scars" (Cliff et al., 1996a) were present in sharks of 138, 143, 144, and 145 cm, including the 131 cm (0 GR) specimen of this study. If the presence of these scars is interpreted as an indication of recent birth, it is tempting to suggest a very wide range in size at birth, between 100 and 145 cm. These scars, however, may persist for some time after birth and given a growth of about 25 cm/yr in newborn sharks, birth size of C. carchirias would be more in the region of that re- ported by Francis (1996), i.e. 92-116 cm (120-150 cm TL). Our range in mass at birth (22-54 kg) is also substantial, which is in keeping with Francis ( 1996 ) who stated that the range in mass at birth for C. carcharias is quite pronounced. Our three mature males had 8, 10, and 13 GRs (293, 317, and 373 cm; 371, 544, and 882 kg, respec- tively). The NSB has caught immature specimens larger and heavier than our smallest mature, e.g. a 295 cm (420 kg) and a 306 cm (442 kg) specimen (Cliff et al., 1996a). Pratt's (1996) smallest mature male was 299 cm (379 cm TL), which is similar to our shark; however, he notes that sizes at sexual matu- rity for male C. carcharias vary widely in the litera- ture. These differences could be due to a variation in length at maturity depending on location, both com- pounded by the use of different length conversion equations and maturity criteria. If the number of GRs is taken as an age estimate, assuming annual GR deposition, male C. carcharias in South Africa would mature between 8-10 years which is similar to the findings of Cailliet et al. (1985) who worked with a size at maturity of 313-365 cm (366-427 cm TL), corresponding to an age of 9-10 years. This age at maturity is higher than that of other lamnids, e.g. that of male I. oxyrinchus at 2-3 years (Pratt and Casey, 1983) and male L. nasiis at about 5 years (Paust and Smith, 1986). No mature female has ever been examined from NSB nets, and the biggest female in our study was 297 cm (6 GRs). The biggest female on record is an immature 348 cm specimen (Cliff et al., 1989), and Bass et al. (1975) reported a mature female of 352 cm (445 cm TL). Again if GRs are deposited annu- ally, the above specimens would be 11 and 12 years, respectively. Age at maturity for female C. carcharias in South Africa would then be at least 12-13 years, slightly higher than that for males. Again, this age at maturity is higher than in other lamnids, e.g. that for female /. oxyrinchus at 7 years (Pratt and Casey, 1983) and female L. nasus at 9-10 years (Paust and Smith, 1986). Although this study could not conclusively prove annual or biannual GR periodicity, the recapture of one shark injected with OTC provided some evidence for annual GR deposition. Assuming that only one growth ring is laid down per year, then C. carcharias from South Africa is relatively slow growing in com- parison with other lamnoids. This observation would support current protective legislation. The NSB nets are now the only directed source of apparent fishing mortality for C. carcharias in South Africa. Cliff et al. ( 1996b ) were of the opinion, using their estimates of mortalities, that the current fishing mortality did not represent overfishing of the white shark stock. Examination of interannual catch rates of C. carcharias in the NSB nets showed an significant decline immediately following the introduction of netting, but thereafter (1978-93) there was no sig- nificant change (Cliff et al., 1996a). Because they are apex predators, C. carcharias are likely to have a small population size (Cliff et al., 1996b), and we have no knowledge about the fecundity and nursery grounds of this species in South African waters. Any possible relaxation of the South African current leg- islation will depend on improved knowledge about population size, immigration and emigration, natu- ral mortality, and fecundity of C. carcharias from South Africa. Any such relaxation, however, is highly unlikely given the increasing protection that has been granted to this species worldwide. It is now also protected in some states of America (Fergusson et al., in press) and Australia (StevensM, where such legislation is based on a more limited knowledge or understand- ing of the biology and papulation dynamics of this species. Acknowledgments The assistance of the field staff of the NSB, who pro- vided specimens and associated information, is greatly appreciated. The laboratory staff was respon- sible for dissecting many sharks and for collecting Wintner and Cliff: Age and growth) determination of Carcharodon carcharlas 167 and storing the vertebrae. We are particularly in- debted to Bortz, Lake & Partners Inc., who provided the equipment to x-ray the vertebrae and especially to W. Stubbs and W. Lake for their time and exper- tise. We would like to thank S. F. J. Dudley for his input and L. J. Natanson for her assistance and valu- able comments on an earlier draft of the manuscript. Literature cited Aasen, O. 1963. Length and growth of the porbeagle iLamna nasus, Bonnaterre) in the north west Atlantic. Fish. Dir Skr Ser. Havunders 13(6):20-37. Bass, A. J., J. D. D' Aubrey, and N. Kistnasamy. 1973. Sharks of the east coast of southern Africa. I: The genus Carcharhinus (Carcharhinidae). Oceanogr. Res. Inst. (Durban) Invest. Rep. 33, 168 p. 1975. Sharks of the east coast of southern Africa. 4. The fami- lies Odontaspididae, Scapanorhynchidae, Isuridae, Cetorhini- dae, Alopiidae, Orectolobidae and Rhiniodontidae. Oceanogr. Res. Inst. (Durban) Invest. Rep. 39. 102 p. Batista, V. S., and T. C. Silva. 1995. Age and growth of juveniles of junteiro shark Car- charhinus porosus in the coast of Maranhao, Brazil. Rev. Brasil. Biol. 55(l):2.5-32. Beamish, R. J., and D. A. Foumier. 1981. A method for comparing the precision of a set of age determinations. Can. J. Fish. Aquat. Sci. 38:982-983. Branstetter, S. 1987. Age and growth estimates for blacktip, Carcharhinus limbatus. and spinner, C. breuipinna, sharks from the northwestern Gulf of Mexico. Copeia 4:964-974. 1990. Early life-history implications of selected carcharhi- noid and lamnoid sharks of the Northwest Atlantic. In H. L. Pratt Jr. S. H. Gruber, and T. Tanmchi (eds). Elas- mobranchs as living resources: advances in the biology, ecology, systematics, and the status of the fisheries, p. 17- 28. U.S. Dep. Commer, NOAA Tech. Rep. NMFS 90. Branstetter, S., and J. A. Musick. 1994. Age and growth estimates for the sand tiger in the Northwestern Atlantic Ocean. Trans. Am. Fish. Soc. 123:242-254. Branstetter, S., J. A. Musick, and J. A. Colvocoresses. 1987. A comparison of the age and growth of the tiger shark. Galeocerdo cuvieri. from off Virginia and from the north- western Gulf of Mexico. Fish. Bull. 85 (2):269-279. Branstetter, S., and R. Stiles. 1987. Age and growth estimates of the bull shark, Carcharhinus leucas, from the northern Gulf of Mexico. Environ. Biol. Fishes 20(3):169-181. Brown, C. A., and S. H. Gruber. 1988. Age assessment of the lemon shark, Negaprion brevir- ostris using tetracycline validated vertebral centra. Copeia 3:747-753. Bruce, B. D. 1992. Preliminary observations on the biology of the white shark, Carcharodon carcharias in South Australian waters. In J. G. Pepperell (ed.), Sharks: biology and fish- eries, p. 1-11. Aust. J. Mar. Freshwater Res. 43. Cailliet, G. M. 1990. Elasmobranch age determination and verification: an updated review. In H. L. Pratt Jr., S. H. Gruber, and T. Taniuchi (eds.), Elasmobranchs as living resources: ad- vances in the biology, ecology, systematics, and the status of the fisheries, p. 157-165. U.S. Dep. Commer.. NOAA Tech. Rep. NMFS 90. Cailliet, G. M., L. K. Martin, J. T. Harvey, D. Kusher, and B. A. Welden. 1983a. Preliminary studies on age and growth of blue, Prionace glauca, common thresher, A/op/as vulpinus, and shortfin make, Isurus oxyrinchus, sharks from California waters. In E. D. Prince and L. M. Pulos (eds.). Proceed- ings of the international workshop on age determination of oceanic pelagic fishes: tunas, billfishes, and sharks, p. 179- 188. US. Dep. Commer. NOAA Tech. Rep. NMFS 8. Cailliet, G. M., L. K. Martin, D. Kusher, P. Wolf, and B. A. Welden. 1983b. Techniques for enhancing vertebral bands in age estimation of California elasmobranchs. In E. D. Prince and L. M. Pulos (eds.). Proceedings of the international workshop on age determination of oceanic pelagic fishes: tunas, billfishes, and sharks, p. 157-165. U.S. Dep. Commer., NOAA Tech. Rep. NMFS 8. Cailliet, G. M., L. J. Natanson, B. A. Welden, and D. A. Ebert. 1985. Preliminary studies on the age and growth of the white shark, Carcharodon carcharias. using vertebral bands. Mem. S. Calif Acad. Sci. 9:49-60. Cailliet, G. M., R. L. Radtke, and B. A. Welden. 1986. Elasmobranch age determination and verification: A review. In T. Uyeno, R. Aral, T Taniuchi, and K. Matsuura (eds.), Indo-Pacific fish biology: proceedings of the second international conference on Indo-Pacific fishes, p. 345- 360. Tokyo: Ichthyological Society Japan. Cailliet, G. M., K. G. Yudin, S. Tanaka, and T. Taniuchi. 1990. Growth characteristics of two populations of Mustelus manazo from Japan based upon cross-readings of verte- bral bands. In H. L. Pratt Jr., S. H. Gruber, and T. Taniuchi (eds.), Elasmobranchs as living resources: ad- vances in the biology, ecology, systematics, and the status of the fisheries, p. 167-176. U.S. Dep. Commer., NOAA Tech. Rep. NMFS 90. Carlander, K. D. 1969. Handbook of freshwater fishery biology, vol. 1, 3rd ed. Iowa State Univ. Press. Ames, lA, 752 p. Casey, J. G., H. L. Pratt Jr., and C. E. Stillwell. 1985. Age and growth of the sandbar shark (Carcharhinus plumbeus) from the western North Atlantic. Can. J. Fish. Aquat. Sci. 42(5):963-975. Chang, W. Y. B. 1982. A statistical method for evaluating the reproducibil- ity of age determination. Can. J. Fish. Aquat. Sci. 39:1208-1210. Cliff, G., S. F. J. Dudley, and B. Davies. 1988. Sharks caught in the protective gill nets ofT Natal, South Africa. l.The sandbar shark Carcharhinus plumbeus (Nardo). S. Afr J. Mar. Sci. 7:255-265. 1989. Sharks caught in the protective gill nets off Natal, South Africa. 2. The great white shark Carcharodon carcharias (Linnaeus). S. Afr J. Mar. Sci. 8:131-144. Cliff, G., S. F. J. Dudley, and M. R. Jury. 1996a. Catches of white sharks in KwaZulu-Natal, South Africa and environmental influences. In A. P. Klimley and D. G. Ainley (eds.). Great white sharks: the biology of Carcharodon carcharias. p. 351-362. Academic Press, Inc., San Diego, CA. Cliff, G., R. P. van der Elst, A. Govender, T. K. Whitun, and E. M. Bullen. 1996b. First estimates of mortality and population size of 168 Fishery Bulletin 97(1), 1999 white sharks on the South African coast. In A. P. Klimley and D. G. Ainley (eds. I, Great white sharks: the biology of Carcharodon carcharias, p. 393-400. Academic Press, Inc., San Diego, CA. Compagno, L. J. V. 1984. FAO species catalogue. Vol. 4: Sharks of the world: an annotated and illustrated catalogue of shark species known to date. Part 2: Carcharhiniformes. FAO Fish. Syn- opsis 125, FAO, Rome, 406 p. 1990. Shark exploitation and conservation, /n H. L. Pratt Jr, S. H. Gruber, and T. Taniuchi (eds.), Elasmobranchs as living resources: advances in the biology, ecology, sys- tematics, and the status of the fisheries, p. 391-414. U.S. Dep. Commer., NOAA Tech. Rep. NMFS 90. 1991. Government protection for the great white shark (Carcharodon carcharias) in South Africa. S. Afr. J. Sci. 87(71:284-285. Corel, Corporation. 1993. Corel DRAW, version 4.0. Corel Corporation, Ottawa, Ontario. Draper, N. R., and H. Smith. 1981 Applied regression analysis, 2nd ed. John Wiley & Sons, New York, NY, 709 p. Ellis, R., and J. E. McCosker. 1991. Great white shark. Stanford Univ. Press, Palo Alto, CA, 270 p. Fergusson, I. K., L. J. V. Compagno, and M. A. Marks. In press. Status account of the great white shark Carcharo- don carcharias (L., 1758) (Chonrichthyes: Lamnidael. In S.L. Fowler i ed. l. Coastal-pelagic species: red list of threat- ened Chondrichthyes. International Union for Conserva- tion of Nature and Natural Resources lIUCN) Species Sur- vival Comm., Shark Specialist Group. Ferreira, B. P., and C. M. Vooren. 1991. Age, growth, and structure of vertebra in the school shark Galeorhinus galeus (Linnaeus, 1758) from southern Brazil. Fish. Bull. 89( 1 1:19-31. Francis, M. P. 1996. Observations on a pregnant white shark with a review of reproductive biology. In A. P. Klimley and D. G. Ainley (eds.). Great white sharks: the biology of Carcharodon car- charias, p. 157-172. Academic Press, Inc., San Diego. CA. Francis, R. I. C. C. 1990. Back-calculation of fish length: a critical review. J. Fish. Biol. 36:88.3-902. Gulland, J. A. 1983. FAOAViley series on food and agriculture. Vol. l:Fish stock assessment: a manual of basic methods. John Wiley & Sons, Chichester. 223 p. Hayashi, J. 1976. Studies on the growth of the red tilefish in the East China Sea -I. A fundamental consideration for age deter- mination from otoliths. Bull. Jpn. Soc. Sci. Fish. 42(11): 1237-1242 Holden, M. J., and M. R. Vince. 1973. Age validation studies on the centra of Raja clavata using tetracycline. J. Con. Int. Explor Mer 35(1):13-17. Kusher, D. I., S. E. Smith, and G. M. Cailliet. 1992. Validated age and growth of the leopard shark. Triakis semifasciata. with comments on reproduction. Environ. Biol. Fish. 35:187-203. McFarlane, G. A., and R. J. Beamish. 1987. Selection of dosages of oxytetracycline for age vali- dation studies. Can. J. Fish. Aquat. Sci. 44:905-909. Mollet, H. F., and G. M. Cailliet. 1996. Using allometry to predict body mass from linear measurements of the white shark. In A. P. Klimley and D. G. Ainley (eds.). Great white sharks: the biology of Carcharodon carcharias, p. 81-89. Academic Press, Inc., San Diego, CA. Mollet, H. F., G. M. Cailliet, A. P. Klimley, D. A. Ebert, A. D. Testi, and L. J. V. Compagno. 1996. A review of length validation methods and protocols to measure large white sharks. In A. P. Klimley and D. G. Ainley (eds.). Great white sharks: the biology of Carcharodon carcharias, p. 91-108. Academic Press, Inc., San Diego, CA. Natanson, L. J., and G. M. Cailliet. 1990. Vertebral growth zone deposition in Pacific angel sharks. Copeia 4:1133-1145. Natanson, L. J., and N. E. Kohler. 1996. A preliminary estimate of age and growth of the dusky shark Carcharhinus obscurus from the south-west Indian Ocean, with comparisons to the western North Atlantic population. S. Afr. .J. Mar Sci. 17:217-224. Parker, H. W., and F. C. Stott. 1965. Age, size and vertebral calcification in the basking shark, Cetorhinus maximus (Gunnerus). Zool. Meded. (Leiden) 40 (34):305-319. Paust, B., and R. Smith. 1986. Salmon shark manual. The development of a com- mercial salmon shark, Lamna ditropis, fishery in the North Pacific. Alaska Sea Grant Rep. 86-01, 430 p. Pratt, H. L., Jr. 1996. Reproduction in the male white shark. In A. P. Klimley and D. G. Ainley (eds. I. Great white sharks: the biology of Carc/iarodo^ carc/ionas, p. 131-138. Academic Press, Inc., San Diego, CA. Pratt, H. L., Jr., and J. G. Casey. 1983. Age and growth of the shortfin mako, Isurus oxyrinchiis, using four methods. Can. J. Fish. Aquat. Sci. 40(11):1944-1957. Punt, A. E., and G. S. Hughes. 1989. PC-YILED II user's guide. Benguela Ecology Programme Report 18, Foundation for Research and De- velopment. Praetoria, South Africa, 60 p. Randall, J. E. 1973. Sizeof the great white shark (Carc/ioroc/ofi). Science (Washington D.C.) 181:169-170. 1987. Refutation of lengths of 11.3, 9.0, and 6.4 m attrib- uted to the white shark. Carcharodon carcharias. Calif Fish Game 73(3):163-168. Ricker, W. E. 1975. Computation and interpretation of biological statistics offish populations. Bull. Fish. Res. Board Can. 191. 382 p. Schwartz, F. J. 19??. Shark ageing methods and age estimation of scalloped hammerhead. Sphyrna lewini, and dusky, Carcharhinus obscurus. sharks based on vertebral ring counts. In E. D. Prince and L. M. Pulos (eds.). Proceedings of the interna- tional workshop on age determination of oceanic pelagic fishes: tunas, billfishes, and sharks, p. 167-174. U.S. Dep. Commer. NOAA Tech. Rep. NMFS 8. Silliman, R. P. 1967. Analog computer models for fish populations. Fish. Bull. 66(1): 31-45. Skomal, G. B. 1990. Age and growth of the blue shark, Prionace glauca, in the North Atlantic. M.S. thesis, Univ. Rhode Island, Kingston, RI, 82 p. Smith, C. L. 1983. Summary of round table discussions on back calcu- Wintner and Cliff; Age and growtfi determination of Carchorodon carcharias 169 lations. In E. D. Prince and L. M. Pulos (eds,), Proceed- ings of the international workshop on age determination of oceanic pelagic fishes: tunas, billfishes, and sharks, p. 45-47. U.S. Dep. Commer., NOAA Tech. Rep. NMFS 8. Smith, J. L. B. 1951. The juvenile of the man-eater, Carcharodon carchar- ias Lmn. Ann. Mag. Hist. Ser 12l4):729-736. STSC (Statistical Graphics Corporation). 1991. Statgraphics, version 5®. Statistical Graphics Cor- poration, Rockville, MD. Ulead Systems, Inc. 1992-94. ImagePals 2 Go!, version 2.0®. Ulead Systems, Inc., Torrance. CA. von Bertalanffy, L. 1938. A quantitative theory of organic growth (Inquiries on growth laws. II). Hum. Biol. 10(2):181-213. Walter, J. P., and D. A. Ebert. 1991. Preliminary estimates of age of the bronze whaler Carcharhinus brachyurus (Chondrichthyes: Carchar- hinidael from southern Africa, with a review of some life history parameters. S. Afr. J. Mar. Sci. 10:37-44. Wintner, S. P., and G. Cliff. 1996. Age and growth determination of the blacktip shark Carcharhinus limbatus from the east coast of South Africa. Fish. Bull. 94:135-144. Yudin, K. G., and G. M. Cailliet. 1990. Age and growth of the gray smoothhound, Mustelus californicus. and the brown smoothhound, M. henlei. sharks from central California. Copeia 1:191-204. Zar, J. H. 1974. Biostatistical analysis, 2nd ed. Prentice-Hall, Inc., Englewood Cliffs, NJ, 718 p. 170 Abstract.— Fish and other animals are often tagged to estimate their abun- dance as well as rates of growth, fish- ing mortality, natural mortality, and movement. Results of these studies are biased if tags are not retained perma- nently and if tag loss is not taken into account. In this paper, we develop a simple tag shedding model to account for the effects of time at liberty, sex, and other factors and use one of its special cases to estimate the instantaneous tag shedding rate from data based on two double-tagging experiments on the school shark. Galeorhinus galeus, and gummy shark, Mustelus antarcticus, off southern Australia. For either species, tag shedding rate could vary with tag type, position of tag on fish, and sex of fish, but not with length at release or time at liberty. The shedding rate of Petersen disc fin tags was well above SO'i'f/yr. Dart tags were shed at a higher rate (41%/yr for school shark; 6.3'7f/yr for gummy shark) than either "Roto" or "Jumbo" fin tags (S'/r/yr for school shark; 6'7f/yr for gummy shark). For either species of shark, the shedding rate of dart tags anchored in the basal cartilage of the dorsal fin was about half that of dart tags anchored in the dorsal musculature. Estimation of instantaneous rates of tag shedding for school shark, Galeorhinus galeus, and gummy shark, Mustelus antarcticus, by conditional likelihood Yongshun Xiao CSIRO Divis)on of Marine Research GPO Box 1538, Hobart, Tasmania 7001, Australia Present address: South Australian Aquatic Sciences Centre SARDI, 2 Hamra Avenue, West Beach, South Australia 5024, Australia E-mail address xiaoyongshun@pi.sa.gov.au Lauren P. Brown Terence I. Walker Marine and Freshwater Resources Institute PO Box 114, Queenscliff, Victona 3225, Australia Andre E. Punt CSIRO Division of Marine Research GPO Box 1538, Hobart, Tasmania 7001, Australia Manuscript accepted 19 March 1998. Fish, Bull. 97:170-184 (1999). Tags are markers placed on or in animals to identify an individual. Animals are tagged to estimate their abundance as well as rates of growth, fishing mortality, natural mortality, and movement. In many studies, a tagged animal is assumed to retain its tag permanently. This assumption, however, is not valid for certain types of tags. Conse- quently, many attempts have been made to estimate tag shedding rates (e,g. Davis and Reid, 1982; Francis, 1989; Faragher and Gordon, 1992; Treble et al,, 1993; Hampton, 1996; Xiao, 1996a). Tag shedding models are of three main tjqjes; all are based on Bever- ton and Holt's (1957, p, 205, equa- tions 14,32-14.37) model for a double-tagging experiment. Some models are conditional on the num- ber of recaptured fish with a single tag, as well as the number of recap- tured fish with both tags as a func- tion of time at liberty, and use the least squares method (GuUand, 1955, 1963; Chapman, 1961; Paulik, 1963; Chapman et al., 1965; Bayliff and Mobrand, 1972; Russell, 1980; Kirkwood, 1981; Alt et al., 1985) or more generally the maximum like- lihood method (Robson and Regier, 1966; Seber, 1973; Seber and Felton, 1981; Wetherall, 1982; Kremers, 1988; Fabrizio et al,, 1996) for esti- mation of parameters. Other mod- els are conditional on the number of recaptured fish retaining at least one tag as a function of time at lib- erty and on the exact times at lib- erty (Wetherall, 1982). Use of these types of models in data analysis re- quires grouping recaptured fish by time at liberty because of an insuf- ficient number of recaptures for a particular (exact) time at liberty, especially in small-scale tagging experiments. Still other models are conditional only on the exact times Xiao et al.: Instantaneous rate of tag shedding for Galeorhinus ga/eus and Mustelus antarcticus 171 at liberty ( Kirkwood and Walker, 1984; Hampton and Kirkwood, 1989; Hearn et al., 1991; Xiao, 1996a). These models 1) use the exact times at liberty in model fitting, 2) use probabilities of tag retention directly rather than using the often statistically un- desirable ratios as the dependent variable in regres- sion analysis, 3) apply to both small (but see below) and large numbers of recaptures, and 4) yield esti- mates of tag shedding rates independent of instan- taneous fishing mortality, natural mortality, and mortalities due to all other causes. Almost all previ- ous tag shedding models have considered only the effects of fish time at liberty on shedding rates, ig- noring effects of other equally or potentially more important factors, such as fish sex and size. School shark Galeorhinus galeus (Linnaeus) and gummy shark Mustelus antarcticus (sensu Last and Stevens, 1994) are major species in the Australian southern shark fishery — a commercial fishery that extends from Western Australia through South Aus- tralia to Bass Strait and Tasmania in the east and that has an annual landed value of $A15.6 million (Walker et al., 1996). Two tagging studies were un- dertaken to study the growth (Moulton et al., 1992), natural mortality (Grant etal., 1979), and local move- ments of these two species within Bass Strait and off eastern Tasmania (T L Walker, Marine and Fresh- water Resources Institute, PO Box 114, Queenscliff, Vic 3225 Australia, unpubl. data ). These studies sug- gest that school shark are highly migratory, compared with gummy shark, but they provide little quantita- tive information about their rates of movements be- yond these areas, where most sharks were tagged and released. Also, fishing effort was too poorly docu- mented at the time of Grant et al.'s (1979) tagging program ( 1940s and 1950s) to be adequate for quan- tifying the rates of movement for these two species. Finally, predominant use of gill nets with large mesh sizes (8 inches) off the southern coast of Western Australia and off South Australia at the time of T.L Walker's tagging study (1970s) led to a low level of fishing effort and a small number of recaptures. Such a lack of quantitative information on rates of move- ment hampered stock assessment. Consequently, a large-scale tagging experiment was designed (Xiao, 1996b) and implemented to fill in this gap. In that study, thousands of individuals were released; each individual was tagged with an easily attachable and highly visible external tag (a Roto tag or a dart tag), the shedding rate of which was to be determined through an accompanying double-tagging experiment (see below). In this paper, we develop a simple tag shedding model to account for the effects offish sex, size, and factors other than time at liberty and use a special case to estimate the instantaneous tag shedding rate for the two species of sharks. Materials and methods Tagging experiments Two double-tagging experiments were performed on G. galeus andM. antarcticus. In the first experiment (Olsen, 1953; Walker, 1989; Table 1), a total of 2597 school and 363 gummy sharks with a respective to- tal length range of 31-164 (85 ±43, «=2586) cm and 32-179 (102 ±24, n=362) cm were captured by long- line hooks, measured to the nearest centimeter, tagged with an internal and external tag, and released in in- shore waters off Victoria, South Australia, and Tas- mania, Australia, fi-om 22 May 1949 to 10 July 1954. Internal tags were either 50 mm long and 23 mm wide (J-tag), or 50 mm long and 22 mm wide (L-tag), or 35 mm long and 10 mm wide (S-tag) and were inserted into the body cavity through an incision on the left flank parallel to the muscles in the lower half of the body immediately below the posterior half of the first dorsal fin. External tags were a white ( W-tag) or gray Petersen disc (G-tag); both were 16 mm in diameter and 1 mm thick and were placed in the midcentral part of the first dorsal fin. Of those released, 417 school and 20 gummy sharks were recaptured within 42.5 years. Their respective total length at recapture ranged from 43 to 175 ( 127 ±35, /!=267 ) cm and from 83 to 152 ( 125 ±19, n = 12) cm; their respective times at liberty ranged from 31 to 15,510 (2761 ±2758, n=417) d, and from 52 to 3900 (1771 ±1159, n=20) d. In the second double-tagging experiment (Table 2), as part of a major tagging experiment (see above), 291 school and 731 gummy sharks with a respective total length range of 38-168 (134 ±17, n=291) cm and 40-176 (108 ±20, n=729) cm were captured in gill nets, measured to the nearest millimeter, tagged with two external tags (a Roto tag and a dart tag) either in the lower half or basal cartilage of the first dorsal fin, and released off southern Australia, from 15 December 1993 to 24 April 1996. Two types of Roto tags were used: either a 45-mm-long and 18-mm-wide Jumbo (Roto) tag, or a 36-mm-long and 9-mm-wide Roto tag (Daltons of New South Wales, AustraHa). The dart tag was 95 mm long and 2 mm in diameter (HallprintofSouth Australia, Australia). As of 1 May 1997, 48 school and 207 gummy sharks were recap- tured. Their respective total length at recapture ranged from 85 to 179 (135 ±18, n=38) cm and from 66 to 167 (115±17,n=150)cm; their respective times at liberty ranged from 31 to 633 (269 ±163, n=48) d, and from 1 to 1138 (275 ±244, a!=207) d. 172 Fishery Bulletin 97(1), 1999 Table 1 Description of the first double-tagging experiment for gummy and school sharks. The num 3er of recaptures include ^, consecu- tive ly and in parentheses, that with two tags. with tag A only, and with tag B only. " — " indicates unknown or not computable. Mean Length Mean Length Mean Range of length at range at length at range at time at time at Number release release Number recapture recapture liberty liberty Row Species Tag A TagB Sex released (cm) (cm) recaptured (cm) (cm) (d) (d) 1 gummy L-tag W-tag M 11 110+07 99-122 — — — — — 2 gummy L-tag W-tag F 1 90 + - 90-090 - - - - — 3 gummy L-tag G-tag M 128 108112 79-144 6(0,6,0) 131 ±16 114-145 2224 ±1154 1209-3900 4 gummy L-tag G-tag F 129 112 ±20 77-179 13(0,13,0) 128 ±15 106-152 1698 ±1104 80-3531 5 gummy S-tag W-tag M 32 86 ±28 33-136 — — — — — 6 gummy S-tag W-tag F 14 65 ±20 38-102 — — — — — 7 gummy S-tag G-tag M 27 88 ±22 39-119 1(0.1.0) 83 ±- 83-083 52 ±- 52-52 8 gummy S-tag G-tag F 21 63 ±25 32-117 — — — — — 9 school J-tag W-tag M 59 127 ±26 62-154 18(2,15,1) 146 + 11 125-155 5039 ±4369 705-15251 10 school J-tag W-tag F 41 128 ±33 60-164 14(1,13,0) 152+15 113-167 3260 ±2333 319-8380 11 school L-tag W-tag M 32 145 ±07 116-160 7(1,6,0) 155 ±14 143-174 4382 ±3142 841-9539 12 school L-tag W-tag F 15 148 ±15 106-160 4(0,4,0) 161 ±08 155-167 3809 ±5548 546-12114 13 school L-tag G-tag — 4 137 ±18 112-155 2(0,2,0) 152 ±- 152-152 2971 ±0769 2427-3515 14 school L-tag G-tag M 521 141 ±12 71-163 127(4,123,0) 147 ±12 114-175 3858 ±3100 82-15510 15 school L-tag G-tag F 292 137 ±17 73-164 71(6,65,0) 149 ±12 112-167 3142 ±2341 89-9107 16 school S-tag W-tag — 2 67 ±09 60-073 — — — — — 17 school S-tag W-tag M 14 48 ±06 40-057 2(0,2,0) 83+- 83-083 2652 ±2456 915^389 18 school S-tag W-tag F 14 54 ±06 43-065 5(0,5,0) 107 +40 57-141 2566 ±1944 260-5262 19 school S-tag G-tag — 15 57 ±12 32-067 2(1,1,0) — — 377 ±0434 70-684 20 school S-tag G-tag M 781 54 ±13 31-148 86(7,79,0) 97+35 43-159 1568 ±1604 31-7555 21 school S-tag G-tag F 807 53 ±12 31-148 79(13,64,2) 95+33 51-159 1221 ±1512 35-6200 Table 2 Description of the second double-tagging experiment for gummy and school sharks. The number of recaptures includes consecu- tively and in parentheses, that with two tags, with tag A only. and with tag B only. "— -" indicates unknown or not computable; | tagging position refers to tag B's position. Mean Length Mean Length Mean Range of length at range at length at range at time at time at Tagging Number release release Number recapture recapture liberty liberty Row Species Tag A TagB position Sex released (cm) (cm) recaptured (cm) (cm) (d) (d) 1 gummy Jumbo dart fin M 68 115 ±08 97-140 13(9,3,1) 117 ±08 107-130 192±119 64-386 2 gummy Jumbo dart fin F 66 125 ±21 80-176 19(17,1,1) 122+10 108-145 119 ±083 13-309 3 gummy Jumbo dart muscle M 101 109 ±14 87-144 41(22,19.0) 112±14 86-148 278 ±232 5-818 4 gummy Jumbo dart muscle F 164 119±21 68-175 43(19,24,0) 123 ±20 91-167 340 ±273 6-1138 5 gummy Roto dart muscle — 1 106 ±- 106-106 1(0,1,0) — — 83 ±- 83-83 6 gummy Roto dart muscle M 151 96 ±18 45-135 37(20,15,2) 106 ±17 66-148 309 ±262 2-10,59 7 gummy Roto dart muscle F 180 99+18 40-136 53(26.26,1) 112+14 78-138 278 +256 1-886 8 school Jumbo dart fin M 46 135+14 108-168 3(3,0,0) 136 ±13 123-149 136 ±089 34-202 9 school Jumbo dart fin F 81 139 ±12 108-167 15(11,3,1) 141 +12 110-157 306 ±130 123-546 10 school Jumbo dart muscle M 77 134+11 100-158 12(9,2.1) 140 ±20 107-179 263 ±201 33-633 11 school Jumbo dart muscle F 53 140 ±14 100-164 9(6,3,0) 142 ±10 122-155 272 ±179 31-551 12 school Roto dart muscle M 13 108 ±23 71-152 3(3,0.0) 110 ±22 85-124 317+164 146-474 13 school Roto dart muscle F 21 110±29 38-160 6(2,4,0) 115 ±15 100-136 229+172 34-468 Xiao et al.: Instantaneous rate of tag shedding for Galeorhinus galeus and Mustelus antarcttcus 1 73 Model Consider a (single) fish i that is captured, double tagged, and released at time t^^d). The index i can be used to examine the effects of any factor on the instantaneous tag shedding rate. Let A and B indicate the two types of tags and P(i,A,B,t(i)) = probability of retaining both tags at time ) = probability that it is caught at time t(i) and reported given that it has retained both tags; C (i,A,0,t(i)) - probability that it is caught at time t(i) and reported given that it has retained only tag A; C (i,0,B,t(i>> - probability that it is caught at time t(i) and reported given that it has retained only tag B; C a, 0,0, t(i)J = probability that it is caught at time t(i) and reported given that it has retained neither tag; U (i,A,B,t(i)) = probability that it is caught at time t(i) but not reported given that it has retained both tags; U (i,A,0,t(i)) = probability that it is caught at time t(i) but not reported given that it has retained only tag A; U (i,0,B,t(i)) = probability that it is caught at time t(i) but not reported given that it has retained only tag B; U (i,0,0,t(i)) = probability that it is caught at time t(i) but not reported given that it has retained neither tag; D (i,A,B,t(i)) = probability that it is dead at time t(i) given that it has retained both tags; D li,A,0,t(i>> = probability that it is dead at time t(i) given that it has retained only tag A; D(i,0,B,t(il) = probability that it is dead at time t(i) given that it has retained only tag B; D(i,0,0,t(i)) = probability that it is dead at time t(i) given that it has retained neither tag; n(i> = probability that it remains alive after type-I mortality (i.e. mortality due to the immediate effects of tagging and handling); pa J) - probability that it retains tag 7 {j=A,B) after type-1 shedding (i.e. tag shedding due to the immediate effects of tagging and handling); F(i,t(i)) = instantaneous rate of fishing mortality at time t{i)\ M(i,t(i>) = instantaneous rate of natural mortality at time t(i); R(i,A,B,t(i)) = probability of reporting given that it is caught at time t(i) and that it has retained both tags R(i,A,0,t; and X(i,B,t(i)) = instantaneous shedding rate of tag B at time <((i. We assume that, in the time interval [t(i),t(i)+At], the probability that fish / retaining both tags is caught is F{i,t(i))AtP{i,A,B,t(i}}+0(M), the probability that it is dead is Mli,t(i))AtP(i,A,B,tli)>+0(At>, the probability that it sheds tag A is X(i,A,t(i))AtP(i,A,B,t(i))+0(At), and the probability that it sheds tag B is X(i,B,t(i)>AtP(i,A,B,t(i))+0(AtJ, where O(At)-^0 as zi^^O. It is also assumed that these events are independent with no more than one event occurring in the time interval. Under these assumptions, the probability that fish i retains both tags at time t(i)+At given that it has retained both tags at time t(i) is given by Pli,A,B,t(i>+At)^[l-F(i,t(i))At-M(i,t(i))At-Mi,A,t(i))At-Mi,B,t(i))AtJP(i,A,B,t(i))+0(At). Taking the limit At^O and letting the dot above a quantity denote the first derivative of that quantity with respect to t(l) yields P(i,A,B,t(i))=-[F(i,t(i))+M(i,t(i» +X(i,A,t(i))+X(i,B,t(i))]P(i,A,B,t(i)). This and similar arguments yield a tag shedding model of the form P{i,A,B,t{i)) = -[F(i,t(i)) + M(i,t(i)) + ?i{i,A,t(i)) + X(i,B,t(i))]P{i,A,B,t{i)) P{i,A.O,t(l)) = -[F(i,t{l)) + M{i,t^i)) + ^i,A,t(i))]P(i.A,0,t(i)) + ^i,B.t(i))P{l,A,B,t^i)) 174 Fishery Bulletin 97(1), 1999 P{i,0,B.t( P{i,0,0,t{ C{i,A,B,t( C(i,A,0,t{ C(i,0,B,t( C{i,O.Oji U(i,A,B,t{ U(i,A,0,t{ U{i,0,B,t( U(i, 0,0, t( D{i,A,B,t{ D{i,A,0,t( D{i,0,B,t( D(i, 0,0, t( )) = -[F{i,t(i)) + M{i,t(i)) + Mi,B,t{i))]P{i,0,B,tti)) + ?i{i,A,t{i))P{i,A,B,tli)) )) = -[F{i,t{i)) + M{i,t{i))]P{i,0,0,t(i)) + Mt,A,tli))P(i,A,0,t{i)) + Mi,B,Hi))P{i,0,B,t(i)) )) = F{i,t{i))R(i,A,B,tU))P(i,A,B,t{i)) )) = F{i,t(i))R(i,A,0,t{i))P{i,A,0,t{i)) )) = F(i,t{i))R{i,0,B,t{i))P{i,0,B,t{i)) )) = F{i,t(i)) = R{i,0,0,fii))P{i,0,0,t(i)) )) = F{i,t(i))[l-R(i,A,B,t{i))]P(i,A,B,t{i)) [to{i)'" p{i,A)e '»'" l-p(/,B)e '°'" tti) - J[F(!,s)+M(i,si]rfs L (III - J-l(i.A,s)ds J fii 1 - JA{i.B.s)ds P{i,0,B,t(i)) = --TTiDe """ l-p(i,A)e '<"■' p{i,B)e '""' - \[F(t,s> + M{i,s)]ds tit) - \Mi.A,s)ds nn - JAli.B,s)ds P{i,0,0,t{i)) = -KiDe ">'" l-p(i,A)e """ l-p{i,B)e """ C{i,A,B,t{i}) = t(i) ■- ' F(i,s}R(i,A,B,s)P(i,A,B,s)ds C{i,A,0,Ui)) = til) F{l,s)Rii,A,0,s)P{i,A,0,s)ds C{i,0,B,t{i)) = Hi) -- ' F(i, s)R(i, 0, B, s)P{i, 0, B, s)ds 'of C(i,0,0,Hi)) = t(i) F(i,s)R{i,0,0,s)P(i,0,0,s)ds U{i,A,B,t(i)) = = ' F{i,s)[l- R(i,A,B,s)]P(i,A,B,s)ds U(i,A,0,t(i]) = /III = ii'(/,s)[l-/?(/,A,0,s)]P(/,A,0,s)ds U{i,0,B,t(i)) = = f(j,s)[l-/?(/,0,B,s)]P(/,0,B,s)c/s [/(/, 0,0, <(/)) = = "f(;,s)[1- i?(/,0,0,s)]P(/,0,0,s)ds 'o<'» D{i,A,B,t{i)) = = M{i,s)P{i,A,B,s)ds 'o<'l D{i,A,0,tU})-- ((J) = ' M{i,s)P(i,A,0,s)ds D(i,0,Bj{i))-- = fM(/,s)P(/;,0,B,s)ds toU) D{i,0,0,t(i))-- Hi) -- ' M{i,s)P(i,0,0,s)d. ? (2) 176 Fishery Bulletin 97(1), 1999 This tag shedding model follows essentially the same line of thought as Xiao's (1996a) and can be readily phrased in the standard terminology of competing risks in survival analysis (David and Moeschberger, 1978). Also, notice that the left-hand side of Equation 1 sums to zero; the left-hand side of Equation 2 sums to n(i). When a single fish is double tagged and released at time t^d), one of 16 mutually exclusive events can happen at time t(i) (Equation 1 or 2). However, only three events are actually observable: the fish has, upon recapture, retained both tags, retained tag A and lost tag B, or lost tag A and retained tag B, with respective probabilities of C(i,A,B,t(i)), C(i,A,0,t(i)) and C(i,0,B,t(i)). The event that it has shed both tags upon recap- ture, with a probability of C(i,0,0,t(i)), cannot be observed, for when both tags are shed, a fish cannot be reliably distinguished from one that was never tagged. A likelihood function can be constructed to estimate parameters in Equation 1 or 2 by following arguments in standard competing risk analysis, but these esti- mates are substantially biased. To overcome this problem, we estimated model parameters by conditioning on observations of three events only, i.e. by maximizing the conditional likelihood function for all reported recap- tures with at least one tag retained i/— aj • ' J-^ ,-) ' -Lj oj with n C(h,A,Bj(h}) \C{h,A,B,t{h}) + C{h,A,0,t,{h)) + C{h,0,B,t(h)} n R{h,A,B,t{h))e(h,A,Bj{h) *-R(h,A,Bj(h))9{h,A,B,t{h)) + R{h,A,0,t(h))e(h,A,0j(h)) + R{h,0,Bjih))9(h,0,B,t{h)) /i = i ^^=n C(j,A,0,tiJ)) J^C(j.A,BMj)) + C{j,A,Oj(j)) + C{j,0,Bj(j) n R{j,A,0,tij))e{j,A,Oj(j)) (3) R{j,A,BjAj))0{j,A.Bjij)) + RU,A,OMjmj^A,O,t(j)) + R{j.O,Bjij))9{j,O,Bj(j)) ^3 0 C(k,0,B.t(k)) \C{k,A.Bjik)) + C{k,A.Oj(k)) + C(k,0,Bj{k)) It n R(k,0,Bj{k})e{k,0,B,t{k) \R{k,A,B,t{k)}9{k,A,B,t(k}) + R{k,A,0,t(k))0{k,A,0,t{k}) + R{k,0,B,t(k))e{k,0,B,t{k)) - \[Hi.A.s\+lii,B.s)]ds d(i,A,B,t{i)) = p{i,A}p(i,B}e '»'" d(i,A,0,t{i)) = p(i,A)e Ja((,A,.s-Ic/s l-p(i,B)e P' ii.B.sUls Xiao et al.: Instantaneous rate of tag shedding for Galeorhinus galeus and Mustelus antarcticus 177 e{i,o,B,t{i)) -1 l-p{i,A)e "•'• Mi,A,s)ds \Mi.B.s)ds e(i,0,0,t{i)) = - jXU,A.s)ds l-p(i,A)e """ p{i,B)e ""• - \X(i.B.s)ds l-p{i,B)e '""' (3) continued where h j, and k index fish recaptures with both tags retained, with tag A only, and with tag B only; n,m , and p are the total numbers of fish recaptures with both tags retained, with tag A only, and with tag B only. In the estimation, we assumed that fg(;J=0, there was no type-I tag shedding (i.e. p{i,A)-pii,B)=l), and R(iA,B,t(i))=R(i,A,Q,t(i))-R(i,0,B.t(i)). The latter assumption makes Equation 3 independent of probability of reporting at time t(i). We also set the instantaneous shedding rate of tagj (;=A,B) as a function offish total length at release L{i) and time at liberty t(i) of the form Mij,t(i))-P^(j}+fi^(j)L(i}+li./j)t(i}, where fi^ij), p^ij) and /J2C/'' are parameters to be estimated. Thus, k(ij,t(i)) has three terms and seven (2^-1) nested models, since each term can be included or excluded in a nested model and a nested model has at least one term. Under these assumptions. Equation 3 becomes ith /-/ — ^ / *-*-'9*A-' O) (4) '-.-n e{h,A,Bj(h)) _\e{h.A,B,t(h)) + 9{h,A,0,t{h)) + d(h,0,B,t(h)) L. n e{j,A,o,t{j) ^/^e(j,A,B,t(j)) + 0(j,A.O,t(j)) + (eij.O,B.t{j)) Ls n e{k.o,Bj[k)) \9{k,A,B,t{k)) + e(k,A,0j(k)) + e{k,0,B,t(k)) 9(i,A,B,t{i)) = e 2 e{i,A,0,t(i)) = e^ ' ^ -[/JolBl+/J,(BlL(il]/((l — /ijiBl/lil 1-e 2 d{i,0,B,t(i)) = e(i,o.oj{!))- -[/ii,(Al+/J,lA)Z,(()]((i) — /JjlAird)- 1-e ^ -[P„iA) + l3^tA)LU)]tU)--p2^-A)tiit 1-e ^ -[0oiB) + l)i{B)Lui\tU)--P2'-B)Ht)'' -|/3„iBl+/3ilB)Z,(i)]((i)-^/32(B)((i) 1-e 2 178 Fishery Bulletin 97(1), 1999 For the first experiment, ?ili,A,t(>))=0 because inter- nal tags (tag A)were inserted into the shark's body cavity and were not shed, except under very unusual circumstances. For the same reason, although three recaptured school sharks appeared to have shed their internal tags (rows 9 and 21, Table 1), these events were actually due to failure to detect the tag upon recapture. Consequently, both tags were assumed to be present for these recaptures. Also, tag shedding rates of white and gray Petersen discs were estimated, singly or in combination, to examine their possible differences (Table 3). Data on X(i,A,t( i>>(Roto tags) Table 3 Instantaneous rate of tag shedding for school shark estimated from data based on the first double-tagging experiment assuming that the shedding rates of internal tags (tag A) are zero, i.e., Mi,A,Hi ))=/3q(A)=0, and those of external tags (tag B depend only on their types, i.e., Mi,B,t(i))=lig(B); n is the number of recaptures. - log(L) gives values of the negative of the logarithm of the | likelihood function; " — " indicates not applicable or not computable . J = J -tag; L = L-tag; S = S-tag; W = W-tag; G = G-tag. The word "and" indicates pooling of data: J and L for pooling data from J-tag and L- tag; M and F for pooling data from males and 1 femal es. Estimates for tag A of J and L and tagB of G are the same as those for tag A of L and tag B of G; estimates for tag A of J and S and tag B of G are the same as those for tag A of 8 and tag B of G. Row Tag A TagB Sex n PgiB) (SE)/yr -log(L) 1 J W MandF 32 0.3718(0.1089) 9.4439 2 J W M 18 0.2829(0.1104) 5.4509 3 J W F 14 0.5816(0.2946) 3.3332 4 L w MandF 11 0.6446(0.3609) 2.3503 5 L w M 7 0.3617(0.2605) 1.3295 6 L w F 4 — — 7 L G Mand F 200 0.7347(0.1012) 45.4817 8 L G — 2 — — 9 L G M 127 1.1439(0.2534) 14.3301 10 L G F 71 0.5202(0.1016) 27.4639 11 S W M and F 7 — — 12 S W — 0 — — 13 s W M 2 — — 14 s W F 5 — — 15 s G MandF 167 3.0653(0.4739) 47.9105 16 s G — 2 1.2692(1.5899) 0.3407 17 s G M 86 4.5992(1.0705) 18.1029 18 s G F 79 2.3.509(0.4955) 26.9553 19 L WandG MandF 211 0.7291(0.0974) 47.8580 20 L W and G — 2 — — 21 L WandG M 134 1.0272(0.2119) 16.8622 22 L WandG F 75 0..5466( 0.1040) 28.3971 23 S WandG M and F 174 3.0857(0.4735) 48.0298 24 s WandG — 2 1.2692(1.5899) 0.3407 25 s WandG M 88 4.5993(1.0702) 18.1029 26 s WandG F 84 2.3912(0.4975) 27.1604 27 J and L W MandF 43 0.4165(0.1084) 12.1642 28 J and L W M 25 0.2993(0.1016) 6.8258 29 J and L w F 18 0.7464(0.3.367) 4.0018 30 J and L WandG M and F 243 0.6460(0.0780) 59.5387 31 J and L WandG — 2 — — 32 J and L W and G M 152 0.7457(0.1255) 27.0143 33 J and L W and G F 89 0.5508(0.0979) 31.7369 34 J and S W M andF 39 0.4434(0.1207) 11.6709 35 J and S W — 0 — — 36 J and S W M 20 0.3162(0.1168) 6.1176 37 JandS W F 19 0.7587(0.3557) 4.4422 38 J and S W and G MandF 206 1.6579(0.21.33) 80.2783 39 J and S W and G — 2 1.2692(1.5899) 0.3407 40 JandS WandG M 106 1.5043(0.2682) 45.5600 41 JandS W and G F 98 1.8729(0.3550) 34.0001 conttnut'd Xiao et al.: Instantaneous rate of tag shedding for Galeorhmus galeus and Mustelus antarcticus 179 were too limited from the second experiment (Table 2) to estimate two or more parameters. We estimated PgiA) only, which can, however, be scaled to p^> were fitted, where possible, to data from each tagging ex- periment. The final and most parsimonious model was decided by the x^ statistic ( Seber and Wild, 1989, p. 196-197). All parameters were estimated by mini- mizing -log(L) by using the simplex algorithm by a FORTRAN 77 program (available on request). Results Maximization of Equation 4 for both sets of tagging data yielded estimates of shedding rate for various (independent) combinations offish sex, tag type, and tag position, and their (asymptotic) standard errors (Tables 3 and 4). If a tag was retained in all recap- tured fish, we assumed that its shedding rate was zero in order to estimate other parameters of the model. Because shedding rates must be nonnegative, the assumption of zero shedding rate will lead to an underestimate of the parameter concerned and in- troduce a positive bias into the estimates of other parameters. The extent of such bias could be assessed by simulation studies but is beyond the scope of this work. Fish length at release or time at liberty, or both, entered certain final models for X(i,B,t(i)), only when the number of fish recaptured was small. By con- trast, whenever there were many fish recaptures (e.g. rows 14-15 and 20-21, Table 1), neither factor en- tered the final model. Therefore, we conclude that fish length at release or time at liberty, or both, did not significantly affect tag shedding rates; and their inclusion in certain models was a result of too few recaptures. Fish sex affected tag shedding rates of Petersen discs for some combinations of tag type and tag position. For a combination of a 50-mm-long and 23-mm-wide inter- nal tag (J-tag) with a white Petersen disc (external) tag (W-tag) (rows 1-3, Table 3), Mi,B,t(i))^0.37l8 (±0.1089)/yr if data are pooled for both sexes of school shark, with a -log-likelihood of 9.4439. For the sex- specific model, A(/,fi,^r/))=0.2829 (±0.1104 )/yr for males; Mi,BMi»=:0.58l6 (±0.2946 )/yr for females, with a (male and female) combined -log-likelihood of 8.7841 (=5.4509-h3.3332). The increase in value of the -log- likelihood function for an extra parameter is, in this case, negligible (x-io2507=2x(9-4439-8.7841)=1.3196), suggesting no statistically significant differences in tag shedding rates between sexes for white Petersen discs. Table 3 (continued) Row Tag A TagB Sex n PgiB) (SE)/yr -\og(L) 42 Lands W MandF 18 0.9094(0.4251) 3.4541 43 Lands W — 0 — — 44 Lands w M 9 0.4791(0.3116) 1.7664 45 L and S w F 9 — — 46 Lands G MandF 367 1.2892(0.1331) 116.1464 47 L and S G — 4 1.2729(1.5769) 0.3409 48 Lands G M 213 2.1071(0.3520) 41.2981 49 Lands G F 150 0.9537(0.1327) 67.8985 50 Lands WandG M andF 385 1.2679(0.1274) 119.8703 51 L and S WandG — 4 1.2729(1.5769) 0.3409 52 L and S WandG M 222 1.8674(0.2992) 45.8682 53 Lands WandG F 159 0.9818(0.1336) 68.9822 54 J and L and S W MandF 50 0.4753(0.1168) 14.1985 55 J and L and S W — 0 — — 56 J and L and S W M 27 0.3252(0.1063) 7.4610 57 J and L and S W F 23 0.8956(0.3763) 4.8981 58 J and L and S G MandF 367 1.2892(0.1331) 116.1464 59 J and L and S G — 4 1.2729(1.5769) 0.3409 60 J and L and S G M 213 2.1071(0.3520) 41.2981 61 J and L and S G F 150 0.9537(0.1327) 67.8985 62 J and L and S WandG MandF 417 1.0891(0.1026) 137.7412 63 J and L and S WandG — 4 1.2729(1.5769) 0.3409 64 J and L and S WandG M 240 1.2738(0.1761) 63.3863 65 J and L and S WandG F 173 0.9478(0.1251) 72.8067 180 Fishery Bulletin 97(1), 1999 Table 4 Instantaneous rate of tag shedding for gummy and school shark < estimated from data based on the second double-taggi ngexperi- ment assuming that A(i, A,Hi))=P„iA} and A((.B,«; ))=/J„lB). Tagg ng position refers to tag B's position; n is the number of recap- tures ; -log(L) gives values of the negative of the logarithm of the likelihood function; " — " indicates not applicable or not comput- able. The word "and" indicates pool ng of data. Jumbo and Roto for pooling data from Jumbo tag and Roto tag; M and F for pooling | data from males and females. Position Row Species Tag A TagB of tag Sex n /3„(A) (SE)/yr jig(B) (SE)/yr -log(Z,) 1 gummy Jumbo dart fm MandF 32 0.1912(0.1349) 0.3770(0.1886) 19.5193 2 gummy Jumbo dart fin M 13 0.2133(0.2125) 0.5642(0.3260) 10.98 3 gummy Jumbo dart fm F 19 0.1771(0.1770) 0.1890(0.1890) 8.0212 4 gummy Jumbo dart muscle MandF 84 — 0.9239(-) 67.5957 5 gummy Jumbo dart muscle M 41 — 0.8550(0.2021) 34.5527 6 gummy Jumbo dart muscle F 43 — 0.9902(-) 32.9376 7 gummy Roto dart muscle M and F 91 0.1304(0.0747) 1.0502(0.1712) 71.5838 8 gummy Roto dart muscle — 1 — — — 9 gummy Roto dart muscle M 37 0.1581(0.1110) 0.8183(0.2187) 31.8327 10 gummy Roto dart muscle F 53 0.0918(0.0913) 1.2111(0.2563) 37.1817 11 gummy Jumbo dart fin and muscle MandF 116 0.0569(0.0402) 0.8278(0,1243) 91.8731 12 gummy Jumbo dart fin and muscle M 54 0.0586(0.0584) 0.8042(0.1749) 47.2285 13 gummy Jumbo dart fin and muscle F 62 0.0555(0.0553) 0.8503(0.1766) 44.6251 14 gummy Jumbo and Roto dart muscle MandF 175 0.0641(0.0369) 0.9828(0.1121) 141.3449 15 gummy Jumbo ind Roto dart muscle — 1 — — — 16 gummy Jumbo and Roto dart muscle M 78 0.0809(0.0570) 0.8379(0.1484) 67.8238 17 gummy Jumbo md Roto dart muscle F 96 0.0447(0.0446) 1.0948(0.1656) 70.9707 18 gummy Jumbo md Roto dart fin and muscle MandF 207 0.0857(0.0381) 0.9172(0.1012) 164.2892 19 gummy Jumbo and Roto dart fin and muscle — 1 — — — 20 gummy Jumbo and Roto dart fin and muscle M 91 0,1011(0.0580) 0.8083(0.1361) 79.4274 21 gummy Jumbo and Roto dart fin and muscle F 115 0.0692(0.0487) 0,9989(0,1474) 82.5257 22 school Jumbo dart fin M and F 18 0.0973(0.0972) 0,2646(0,1530) 11.0858 23 school Jumbo dart fin M 3 — — — 24 school Jumbo dart fin F 15 0.1104(0.1103) 0.2948(0.1706) 10.6735 25 school Jumbo dart muscle M and F 21 0.1041(0.1038) 0.4262(0.1917) 14.7690 26 school Jumbo dart muscle M 12 0.1727(0.1725) 0.3219(0.2282) 6.6030 27 school Jumbo dart muscle F 9 — 0.5484(0.3201) 7.3704 28 school Roto dart muscle MandF 9 — 0.7845(0.3967) 8.5426 29 school Roto dart muscle M 3 — — — 30 school Roto dart muscle F 6 — 1.6867(0.8865) 5.6360 31 school Jumbo dart fin and muscle MandF 39 0,1003(0.0708) 0.3466(0.1230) 26.0748 32 school Jumbo dart fin and muscle M 15 0.1421(0.1419) 0.2700(0.1912) 7.0831 33 school Jumbo dart fin and muscle F 24 0.0773(0.0772) 0.3831(0.1571) 18.7670 34 school Jumbo and Roto dart muscle Mand F 30 0.0798(0.0796) 0.5339(0.1793) 24.0789 35 school Jumbo and Roto dart muscle M 15 0.1188(0.1188) 0.2263(0.1602) 7.5881 36 school Jumbo and Roto dart muscle F 15 — 0.8882(0,3425) 14.0341 37 school Jumbo and Roto dart fin and muscle Mand F 48 0.0876(0.0619) 0.4254(0,1233) 35.8236 38 school Jumbo and Roto dart fin and muscle M 18 0,1038(0,1037) 0.1997(0.1414) 7.9368 39 school Jumbo and Roto dart fin and muscle F 30 0.0746(0.0745) 0.5510(0.1755) 26.7306 However, for a combination of a 50-mni-long and 22- mm-wide internal tag (L-tag) with a gray Petersen disc (external) tag (G-tag) (rows 7-10, Table 3), A(i,B.t(i))= 0.7347 (±0.10r2)/yr if data are pooled for both sexes, with a -log-likelihood of 45.4817. For the sex-specific model, ?i(i,B,t(i)) = l.l439 (±0.2534)/yr for males; ?i(i,B,t(i))=0.5202 (±0.1016)/yr for females, with a (male and female) combined -log-likelihood of 41.7940 (=14.3301+27.4639). The increase in value of the -log- likelihood function for an extra parameter is statisti- cally significant (x'l oo()66=2x<45. 4817-41. 7940)= 7.3754), suggesting significant differences in tag shed- ding rates between sexes for gray Petersen discs. Simi- larly, for a combination of a 35-mm-long and 10-mm- wide internal tag (S-tag) with a gray Petersen disc (ex- ternal) tag (rows 15-18, Table 3), A(;,B,^(/;i=3.0653 (+0.4739)/vr if data are pooled for both sexes, with a -log-likelihood of 47.9105. For the sex-specific model, Xd.Bjd ))^4.5992 (±1.0705)/yr for males; k(i,B,tii>)= 2.3509 (±0.4955 )/yr for females, with a (male and fe- Xiao et al.: Instantaneous rate of tag shedding for Galeorhinus galeus and Mustelus antarcticus 181 male) combined -log-likelihood of 45.0582 ( = 18.1029+ 26.9553). The increase in value of the -log-likelihood function for an extra parameter is, again, statistically significant (x2iooi69=2x*47. 9105-45.0582)^5. 7046), again suggesting significant differences in tag shed- ding rates between sexes for gray Petersen discs. No- tice, in these cases, that tag shedding rates for males nearly doubled those for females. For the second tag- ging experiment, no differences in tag shedding rates were found among sexes for either species of shark (Table 4). The shedding rate of Petersen discs for the school shark was very high. When combined with a 50-mm- long and 23-mm-wide internal tag (J-tag), white Petersen disc (W-tag) had a shedding rate of Af(,B,^fiiJ=0.2829(±0.1104)/yr or 100x(l-e-0 2829) = 24.64%/yr for males, and A('/,fi,;f/J)=0.5816{+0.2946)/ yr or 44.10%/yr for females (rows 1-3, Table 3). When combined with a 50-mm-long and 22-mm-wide inter- nal tag (L-tag), gray Petersen disc (G-tag) had a shed- ding rate of Af/,S,mi)= 1. 1439 (±0.2534 )/yr or 68. 14%/ yr for males and Af/,B,) in the like- lihood function (Equation 3 or 4) are cancelled out. Thus, as in Xiao (1996a), our tag shedding model applies, even when F(i,t(i)) and M(i,t(i)) are arbitrary functions of time t(i). On the other hand, if fishing and natural mortalities depend on the state variables of tags A and B, then terms in P(i,A,B,t(i)), P(iA,0,t(i)),P(i,0,B,t(i» and P(i,0,0,t(i))invo\ying{ouT fishing mortalities F(i,A,B,t(i)), F(i,A,0,t(i)), 7^f;,0,B,^fiiJ and Ff(,0,0,«/Ji and four natural mortali- ties M(i,A,B,t(i)), M(i,A.O,t(i>), M{i,0,B,t(i)) and M(i,0,0,t(i)) cannot be factored out. Then, for esti- mation of parameters by maximizing Equation 3, particular functional forms of all the eight mortali- ties must be hypothesized. This tag shedding model is more general but more data-demanding. The other interesting feature of our tag shedding model is that Equation 3 is independent of probabilities of report- ing R(i,A,B,t(i)), R(i,A,0,t(i)), R(i,0,B,t(i)) and R(i,0,0,t(i)) if these probabilities are identical, arbi- trary functions of time t(i) because of the way they enter Equation 3. Statistically significant differences in shedding rates of Petersen discs between male and female school sharks were detected when many fish were recaptured. We do not know why such differences existed but we postulate that male sharks have a higher tag shedding rate because they are more ac- tive and would tend to rub off the tags and that fe- male sharks have a lower tag shedding rate because they are larger and have thicker fins. An external fin tag, such as a Petersen disc, is shed only after its pin or locking mechanism has cut through the fin. The larger the tagged fish, the thicker is its fin and hence the farther the distance its pin or locking mechanism has to cut through to the posterior edge of the fin. Consequently, larger animals have lower shedding rates. Thus, sex is confounded in its effects with size. That is probably why the length at release of school sharks did not affect the shedding rates of Petersen discs within a wide size range examined, although the loss of anchor tags (Floy tags) was size- dependent for striped bass Morone saxatilis ( Waldman et al., 1990 ) but size-independent for lake trout Salvelinus namaycush (Fabrizio et al., 1996). We could not detect differences between sexes with fewer recaptures, however, because the use of Equa- tion 1 or 2 to resolve sexual differences in tag shed- ding rate requires many recaptures (see below). Shedding rates of Petersen discs, Roto tags, and dart tags did not change with time at liberty. Some 182 Fishery Bulletin 97(1), 1999 tagged fish have higher shedding rates than others, because tags that are less securely attached are shed earlier. The proportion of less securely attached tags decreases with increasing time at liberty. This will yield an apparent decrease in tag shedding rate with time at liberty. A similar argument applies when tag shedding rates vary among individuals. The lack of a trend may indicate negligible tag losses ft-om im- proper attachment, insignificant individual variability in tag shedding rate, or insufficient data (see below). Estimates of tag shedding rates in Tables 3 and 4 must be used cautiously because only those that are based on many recaptures are reliable, whereas those that are based on few recaptures are unreliable. For example, the estimates of tag shedding rates for a combination of a 50-mm-long and 22-mm-wide in- ternal tag (L-tag) with a white Petersen disc (W-tag, external) (rows 4-6, Table 3) were based on only 11 recaptures (rows 11 and 12, Table 1), only one of which had retained both tags (row 11, Table 1), and hence are unreliable. No estimates could even be obtained for a combination of a 35-mm-long and 10- mm-wide (S-tag) internal tag with a white Petersen disc (W-tag, external) (rows 11-14, Table 3), despite seven recaptures, none of which had retained both tags (rows 16-18. Table 1). Similarly, no estimates could be obtained, for any tag combinations, from data on gummy sharks from the first double-tagging experiment, despite 20 recaptures, none of which had retained both tags (rows 1-8, Table 1). Equally un- reliable estimates of tag shedding rates could also result from pooling of information while ignoring differences in its sources. For example, estimates from pooling all three internal tags (i.e. J-tag, L-tag and S-tag) (rows 54-65, Table 3) should be treated cautiously because of sexual differences inferred above. By contrast, for both sexes of school sharks, the estimates of shedding rates of gray Petersen discs are reliable for its combination with a 50-mm-long and 22-mm-wide internal tag (L-tag) (rows 9 and 10, Table 3) or with a 35-mm-long and 10-mm-wide (S- tag) internal tag (rows 17 and 18, Table 3) because information from many fish recaptures was used in their estimation. Much less reliable estimates were obtained for dart tags on gummy sharks (rows 5, 6, 9, and 10, Table 4 ). Although rather high in all cases, all these shedding rates are actually underestimated, as will be shown and published elsewhere. Although we have examined only the effects of tag type, sex, length at release, and time at liberty on tag shedding, many other factors, such as tagging operator (Hampton, 1996), can also affect tag shed- ding rate. However, hundreds or even thousands of fish need to be recaptured (many more need to be released) to estimate effects of tagging operators re- liably. Such a great demand of data is well expected of Equation 1 or 2, which is a compartmental model. The solution of a compartmental model can be given by a linear combination of exponentials and is known to yield bad ill-conditioning (Seber and Wild, 1989, p. 118-119). Indeed, for some compartmental mod- els, no amount of data is sufficient for identifjdng model parameters. Similarly, the "best" model of all possible models of a general model is identifiable only by a sufficient volume of data. As mentioned above, fish length at release or time at liberty, or both, en- tered certain "best" models for X(i,B,t(i)), when the number of fish recaptured was small, but did not, when there were many fish recaptures. This finding suggests that fewer data than sufficient cannot iden- tify the "best" model. To detect and address problems with parameter and model identifiability for a par- ticular general model ( e.g. Equation 1 or 2 ), one might generate as large a set of data as necessary, for ex- ample, by duplicating each record of an existing set of data from a double-tagging experiment a neces- sary number of times, analyse it, and design one's tagging experiment accordingly (e.g. to determine the number offish to be released and the expected num- ber offish to be recaptured). Results of our study have major implications for future double-tagging experiments for estimating instantaneous tag shedding rate and for analysis of tagging data. Because estimation of a single para- meter requires many fish recaptures and hence in- curs considerable financial resources, use of an eas- ily detected and permanent tag eliminates a need for considering tag loss and is preferred in any tag- ging experiment. However, with a commercially or recreationally harvested species, problems of tag re- porting remain. Use of two readily detectable, identi- cal tags with a moderate shedding rate in a double- tagging experiment reduces the number of parameters to be estimated by one half A moderate shedding rate is necessary because too low a shedding rate requires some recaptures after a long time at liberty for reliable estimation of parameters; too high a shedding rate ren- ders the tag useless for some applications. Acknowledgments We wish to thank Mick Olsen of the CSIRO Division of Fisheries for collecting and making available to us data for the first tagging experiment. We also thank Natalie F. Bridge (Victorian Marine and Fresh- water Resources Institute) for her field work and for managing the data, and Grant West and John D. Stevens (CSIRO Division of Marine Research) for their help with the tagging data and practicalities of Xiao et al : Instantaneous rate of tag shedding for Galeorhinus galeus and Mustelus antarcticus 183 both tagging experiments. We also sincerely thank Shuichi Kitada and an anonymous referee for their constructive comments. The work was funded by the Australian Fisheries Management Authority and Fisheries Research and Development Corporation. This is SharkFAG document SS/97/D8. Literature cited Alt, G. L., C. R. McLaughlin, and K. H. Pollock. 1985. Ear tag loss by black bears in Pennsylvania. J. Wildl. Manage. 49:316-320. Bayliff, W. H., and L. M. Mobrand. 1972. Estimates of the rates of shedding of dart tags from yellowfin tuna. Inter-Am. Trop. Tuna Comm. Bull. 15:441-462. Beverton, R. J. H., and S. J. Holt. 1957. On the dynamics of exploited fish populations. Fish. Invest. Ser. 2, Mar. Fish. GB Minist. Agric. Fish. Food 19, 533 p. Chapman, D. G. 1961. Statistical problems in dynamics of exploited fisher- ies populations. Proc. Berkeley Symp. Math. Stat. Probab. 4:153-168. Chapman, D. G., B. D. Fink, and E. B. Bennett. 1965. A method for estimating the rate of shedding of tags from yellowfin tuna. Inter-Am. Trop. Tuna Comm. Bull. 10:335-352. David, H. A., and M. L. Moeschberger. 1978. The theory of competing risks. Griffin's Statistical Monographs and Courses No. 39, Alden Press, Oxford, 103 p. Davis, T. L. C, and D. D. Reid. 1982. Estimates of tag shedding rates for Floy FT-2 Dart and FD-67 Anchor Tags in barramundi, Lates calcarifer (Blochl. Aust. J. Mar Freshwater Res. 33:1113-1117. Fabrizio, M. C, B. L. Swanson, S. T. Schram, and M. H. Hoff. 1996. Comparison of three nonlinear models to describe long-term tag shedding by lake trout. Trans. Am. Fish. Soc. 125:261-273. Faragher, R. A., and G. N. G. Gordon. 1992. Comparative exploitation by recreational anglers of brown trout, Salmo trutta L., and rainbow trout, Oncorhynchus mykiss (Walbaum), in Lake Eucumbene, New South Wales. Aust. J. Mar Freshwater Res. 43:835-845. Francis, M. P. 1989. Exploitation rates of rig iMustetus lenticulatus) around the South Island of New Zealand. N.Z. J. Mar. Freshwater Res. 23:239-245. Grant, C. J., R. L. Sandland, and A. M. Olsen. 1979. Estimation of growth, mortality and yield per recruit of the Australian school shark, Galeorhinus australis (Macleay), from tag recoveries. Aust. J. Mar Freshwater Res. 30:625-637. Gulland, J. A. 1955. On the estimation of population parameters from marked members. Biometrika 42:269-270. 1963. On the analysis of double-tagging experiments. Int. Comm. Northwest Atl. Fish. Spec. Publ. 4:228-229. Hampton, J. 1996. Estimates of tag-reporting and tag-shedding rates in a large-scale tuna tagging experiment in the western tropi- cal Pacific Ocean. Fish. Bull. 95:68-79. Hampton, J., and G. P. Kirkwood. 1989. Tag shedding by southern bluefin tuna Thunnus maccoyii. Fish. Bull. 88:313-321. Heam, W. S., G. M. Leigh, and R. J .H. Beverton. 1991. An examination of a tag-shedding assumption, with application to southern bluefin tuna. ICES J. Mar. Sci. 48:41-51. Kirkwood, G. P. 1981. Generalized models for the estimation of rates of tag shedding by southern bluefin tuna ( Thunnus maccoyii). J. Cons. Int. Explor. Mer 39:256-260. Kirkwood, G. P., and M. H. Walker. 1984. Anew method for estimating tag shedding rates, with application to data for Australian salmon, Arripis trutta esper Whitley Aust. J. Mar Freshwater Res. 35:601-606. Kremers, W. K. 1988. Estimation of survival rates from a mark-recapture study with tag loss. Biometrics 44:117-130. Last, P. R., and J. D. Stevens. 1994. Sharks and rays of Australia. CSIRO, Australia, 513 p., with 84 plates. Moulton, P. M., T. I . Walker, and S. R. Saddlier. 1992. Age and growth studies of gummy shark, Mustelus antarcticus Guenther, and school shark, Galeorhinus galeus (Linnaeus), from southern-Australian waters. Aust. J. Mar Freshwater Res. 43:1241-1267. Olsen, A. M. 1953. Tagging of school shark, Galeorhinus australis (Macleay) (Carcharhinidae), in south-eastern Australian waters. Aust. J. Mar. Freshwater Res. 4:95-104. Paulik, G. J. 1963. Estimates of mortality rates from tag recoveries. Biometrics 19:28-57. Robson, D. S., and H. A. Regier. 1966. Estimates of tag loss from recoveries offish tagged and permanently marked. Trans. Am. Fish. Soc. 95:56-59. Russell, H. J. 1980. Analysis of double-tagging experiments: an update. Can. J. Fish. Aquat. Sci. 37:114-116. Seber, G. A. F. 1973. The estimation of animal abundance and related parameters. Griffin. London, 506 p. Seber, G. A. F., and R. Felton. 1981. Tag loss and the Petersen mark-recapture experi- ment. Biometrika 68:211-219. Seber, G. A. F., and C. J. Wild. 1989. Nonlinear regression. John Wilev& Sons. New York, NY, 768 p. Treble, R. J., R. W. Day, and T. J. Quinn II. 1993. Detection and effects on mortality estimates of changes in tag loss. Can. J. Fish. Aquat. Sci. 50:1435- 1441. Waldman, J. R., D. J. Dunning, and M. T. Mattson. 1990. A morphological explanation for size-dependent an- chor tag loss from striped bass. Trans. Am. Fish. Soc. 119:920-923. Walker, T. I. 1989. Methods of tagging adopted in the southern shark fishery. In DA. Hancock (ed). Bureau of Rural Resources, proceedings no. 5: Australian Society for Fish Biology Tag- ging workshop; Sydney 21-22 July 1988), p. 105-108. Aus- tralian Government Pubhshing Service, Canberra. Walker, T., T. Stone, T. Battaglene, and K. McLoughlin. 1996. Fishery assessment report: the southern shark fish- ery 1995. Victorian Marine and Freshwater Resources Institute, Australia, 55 p. 184 Fishery Bulletin 97(1), 1999 Wetherall, J. A. at liberty for a double tagging experiment. Can. J. Fish. 1982. Analysis of double-tagging experiments. Fish. Bull. Aquat. Sci. 53:1852-1861. 80:687-701. 1996b. A framework for evaluating experimental designs Xiao, Y. for estimating rates offish movement from tag recoveries. 1996a. A general model for estimatmg tag-specific shed- Can. J. Fish. Aquat. Sci. 53:1272-1280. ding rates and tag interactions from exact or pooled times 185 Areas, depths, and times of high discard rates of scup, Stenotomus chrysops, during demersal fish trawling off the northeastern United States Steven J. Kennelly Manomel Center for Conservation Sciences PO. Box 1770, Manomet, Massachusetts 02345 Present address: NSW Fishenes Research Institute PO. Box 21, Cronulla 2230, Australia E-mail address, kennellsigfisheries. nsw.gov.au In the waters off the northeastern United States, many stocks of com- mercial and recreational species have declined in recent years (An- thony, 1993; Sissenwine and Rosen- berg, 1994; Collins, 1995; Ma- tthiessen, 1995; NEFSC, 1995). Al- though much of the blame for these declines is ascribed to sustained overfishing, there also has been substantial concern over bycatch and discarding practices in key fisheries of the region, especially those involving demersal trawling (Murawski, 1994; Howell and Langan, 1987, 1992; Cadrin et al., 1995; Kennelly et al., 1997). For many years scup (or porgy, Stenotomus chrysops) has been an important commercial species in the mid-Atlantic and southern New England regions, caught princi- pally by otter trawling and to a lesser extent by pound nets, float- ing trap nets, and fish traps (Shep- herd and Terceiro, 1994). Like many other key species in the re- gion, annual commercial landings of this species have declined mark- edly in recent years from between 18,000 and 27,000 metric tons (tl in the 1950s and 60s to 6000 t in 1992 and 4400 t in 1993 (NEFSC, 1995). There is also an important recreational fishery for scup in this region, and recreational landings in recent years have accounted for 20 to 50'7f of the total annual catch. These have also declined from an estimated 3,100 t in the 1980s to 2100 t in 1992 and 1300 t in 1993 (NEFSC, 1995). Scientists at the 1995 Northeast Regional Stock As- sessment Workshop for the scup stock in this region concluded that 1) it is currently overexploited; 2) it is at a low level of biomass; and 3 ) current high rates of exploitation of age 0-2 fish should be decreased as much as possible (NEFSC^ ). As predicted by Wilk and Brown (1980) some time ago, one of the causes of mortality for young scup in this region is thought to be the incidental bycatch and subsequent discarding of this species from de- mersal trawlers that target other species, particularly squid (Loligo spp.). McKiernan and Pierce's^ study of the inshore squid fishery in Nantucket and Vineyard Sounds, Massachusetts, showed significant discard of scup but, like Cadrin et al. (1995), they concluded that the inshore abundances of this species were more probably related to trawl effort throughout the region than to the relatively small effort of inshore squid trawlers. They concluded that significant numbers of small scup may be discarded by squid trawlers farther offshore where scup are known to migrate in the autumn (see also Finkelstein, 1971; Eklund and Targett, 1990) and recom- mended an examination of the dis- carding practices of these trawlers. The most reliable way to quan- tify discards in commercial fisher- ies is for observers to record data during normal fishing operations (e.g. Jean, 1963; Powles, 1969; Young and Romero, 1979; Atkinson, 1984; Murawski et al., 1995). Infor- mation from such programs is a necessary prerequisite for the two main management alternatives used to reduce discards: 1) spatial and temporal closures to fishing in areas and times of high rates of dis- card of key species (i.e. discard "hot- spots") (Murawski, 1992; Hendrick- son and Griffin, 1993; Alverson et al., 1994, Kennelly, 1997); and 2) modifications to fishing gears and practices that improve selectivity (Robertson and Stewart, 1988; Isaksen et al., 1992; Hall, 1994; Broadhurst et al., 1996; Broadhurst and Kennelly, 1997). Since 1989, the National Marine Fisheries Service's Northeast Fish- eries Science Center (NEFSC) has operated a large-scale observer pro- gram in many of the fisheries off the northeastern United States from Maine to North Carolina (Murawski etal., 1995; Kennelly et al., 1997). The data collected from demersal trawlers in this program have provided an opportunity to examine the spatial and temporal occurrences of discarded scup in the offshore mid-Atlantic and southern New England trawl fishery. ' NEFSC (Northeast Fisheries Science Center). 1995. Report of the 19th northeast regional stock assessment work- shop-the plenary. NOAA/NMFS/NEFSC, Woods Hole, MA. NEFSC Ref Doc. 95-09, 57 p. - McKiernan, D. J., and D. E. Pierce. 1995. Loligo squid fishery in Nantucket and Vineyard Sounds. Massachusetts Div. Fish. Publ. 17648-75-200-1/95-3.47- CR, 62 p. Manuscript accepted 7 April 1998. Fish. Bull. 97:185-192 (1999). 186 Fishery Bulletin 97(1), 1999 Materials and methods Field sampling Each month between July 1990 and June 1994, Manomet Center for Conservation Sciences (under contract to the NEFSC) placed observers on board demersal trawlers working throughout the northeast- ern United States. The selection of vessels and trips undertaken did not adhere to any consistent survey design but was based on changing national, man- agement and stock assessment priorities (Murawski et al., 1995) in addition to varying concentrations of fishing fleets in particular areas and times. As a con- sequence, the observer coverage of these trawlers was highly uneven throughout the period; some areas and times received high sampling and others received very little or no sampling. Of relevance to this paper are the observer data gathered in the southern New England and mid-Atlantic areas shown in Figure 1. During observed trips, the contents of the codend from each tow were emptied onto the deck and sorted by the crew. If the entire catch could not be sampled, data were collected from a representative subsample. The data collected from each tow that were of rel- evance to this paper were the following; area, depth, date, time and tow duration, the weights of retained and discarded individuals of various species in the catch, the number of species caught, and the total weights of all species retained or discarded. Because of various operational and logistic reasons, not all tows from all trips were sampled for the weights of retained and discarded species; however, the data examined in this paper include only those tows that were completely sampled. Analysis of data Because of the lack of a consistent survey design in this observer program, the data generated could not be regarded as randomly collected, independent samples of the trawling effort of the region, and thus violated the basic assumptions required for conven- tional statistical analyses (see also Kennelly et al., 1997). Nevertheless, because of the size of the data set and its extensive spatial and temporal coverage, large subsets of data could be extracted for many areas and months and therefore permitted the iden- tification of certain key areas, depths, and times of consistently high rates of scup discard (see "Results" section). The first and broadest examination of the data was to plot the discard rates for scup for all the areas sampled in the region to identify those areas that experienced consistently high rates. Next, dis- card rates for scup from each sampled tow in each of these identified areas was plotted against depth to 80" w 76" W , , I.,.,.. 74" W , ,.J., . , 72" W 70" W 68" W 66 W 4-— 40 N - Long Island »JLy;_r,61, New York City V. ^(^ K'M 38° N Chesapeake Bay ; v . ■ ■^ 36° N - Cape Hatteras T^ 50f 500f Figure 1 Map showing the various NMFS statistical areas examined in the present paper Areas labelled in bold are those with sufTicient observer coverage to be examined in Figure 2. NOTE Kennelly: Discard rates for Stenotomus chrysops 187 determine which depths yielded consistently high rates. Once these depths were identified, the dis- card rates of scup in these tows were plotted against time to determine the particular months when the high discards occurred. In this way, the large observer data set for the region was fo- cussed to identify the particular areas, depths, and months that yielded high discard rates for scup. Finally, catches of other common species in these identified areas and times were examined by plotting their retention and discard rates. Results Figure 2 shows the aver- age discard rates for scup (in pounds per hour towed) for the areas de- fined in Figure 1. All ar- eas other than 526, 537, 539, 613, 615, 616, 621, 622, and 623 showed low discard rates for scup (defined as a maximum rate of less than 20 Ib/h) or low levels of observer coverage (defined as less than 10 observed tows during the study), or both. Figure 3 shows discard rates for scup against the depth of the tows sampled in each of the areas iden- tified in Figure 2 above. Most areas recorded high discard rates across a range of depths, except areas 613 and 621 which showed marked concen- trations from 30^0 fath- oms (fm) and 6-17 fm respectively. o o o ■^ Number of tows observed o o o o 1 ^ 8 0 o moo CM T- ,- in o o o o o o o O in o Q oj »- T- iJS o o o 8 ^ 2 S . a o I o o . 2 8 (3S) P9«oi jnog jad papjeosjp dnos jo spunod uea^ ■a 3 en _o a 3 o T3 -a a, 3 188 Fishery Bulletin 97(1), 1999 500- 400- 300- 200- 100- ^ 0- 5 o 1 2,000 -, D- 1,500 -a 1 300 1 S. 250 200 150 100 50 0 C Scup disca discard. 526 o 4,000 1 3,500 3,000 2,500 H 2,000 1,500 1 ,000 4 o_ e ^ 500 1 537 o 800 T 700 o 600 o 500 400 -fe 300 °o^ 200 613 621 600t 0 500-1 0 400 „ 0 cP oS> 300 ^ o 200^0 - -'^- "^Mffri n 0 ^ " ° 615 0 3,500 1 3,000 2,500 2,000 1,500 1 0 1,000 500 J 622 o 539 o 6,000 1 5,000 4,000 o ^ ° 3,000 g °o 2,000 4^ 1.000 616 o 12,000- ^ 10,000- 8,000 - 6,000 - CO ° o 4,000- \..^l3 oo 2,000- 623 o o 10 20 30 40 50 60 70 80 90 100 0 10 20 30 40 50 60 70 80 90 100 0 10 20 30 40 50 60 70 80 90100 Depth (fm) Figure 3 rd rates (per hour trawled) plotted against depth for those areas in Figure 2 that showed consistently high rates of scup Figure 4 shows the timing of discard rates for scup in area 613 between 30 and 40 fm and in area 621 between 6 and 17 fm. For area 613, the highest rates at those depths occurred in November and December of each of the four years studied. Such a persistent trend for high scup discards was not evident for area 621, where highest rates occurred in May, August, and Sep- tember 1991, and in September and October 1992. Figure 5 shows the mean number of pounds re- tained and discarded for the most common species caught in all tows sampled in area 613, between 30 and 40 fm, in the months of November and Decem- ber each year. Because of inconsistent identifications, data for squid and skates include data for all such species. Squid were by far the main retained species with an average of 507 Ib/h. Scup and whiting, Merluccius bilinearis, were also retained in signifi- cant quantities at 227 and 223 Ib/h respectively. The main discarded species was dogfish, Sqiioliin acanthios, (672 Ib/h), followed by scup i319 Ib/h) and skates (236 Ib/h), and lesser quantities of butterfish, Peprilua triacanthus, (96 Ib/h), whiting ( 72 Ib/h), and red hake, Urophycis chuss, (51 Ib/h). Discussion The above treatment of a four-year subset of the NMFS sea sampling database indicated that trawl- ing in a particular area (area 613), depth (30-40 fm), and time of year (November to December), consis- tently led to high discard rates for scup. Before dis- cussing this result, however, it is necessary to con- sider the problems inherent in the data analyzed. In particular, the design of this observer program in- volved a nonrandom, nonindependent allocation of sampling effort that was uneven in space and time (see Fig. 2). These problems precluded 1) the use of conventional statistical analyses to detect trends and 2) the identification of patterns for all areas for all months. For example, the uneven observer coverage may have resulted in the identification of only 9 of the 20 statistical areas in the region as having high discard rates for scup (Fig. 2). Although adequate obsei^ver coverage seemed to occur across most of the depths in many of these nine areas (Fig. 3), not all areas had all depths sampled, precluding the con- clusion that other depth-related areas of high scup NOTE Kennelly: Discard rates for Stenotomus chrysops 189 3,000 Area 613, 30-40ttm 2,500 o ° o 2,000 o ^ o° 1 1.500 &> o 1 1.000 ° %° .c e e 0 S 500 O R ° ° §§ a £>.£a8§i> 1 Q. -^" ■D i 1 1 1 1 1 1 1 1 — 0 — ^ — Q 1 1 1 1 — Q 1 1 CO « 800 TJ Area621.6-17ftm CL ''00- o 13 g 600- o 500 o w pun o ° 300 8 200 0 o 8 „ o o J A S O N D JFMAMJJASOND JFMAMJJASOND JFMAMJJASOND J F M A M J 1990 1991 1992 1993 1994 Figure 4 Scup discard rates (per hour trawled) plotted against time for those depths and areas in Figure 3 that showed consistently high | rates for scup discard. discard did not exist in the region (i.e. in addition to the identified depths in areas 613 and 621). Figure 4 shows the temporal pattern of scup discard in those areas and depths identified from Figures 2 and 3 and reveals a consis- tent pattern of high discard for area 613 but not for area 621. However, more observer coverage throughout the year could have provided evidence of other consistent peaks of scup discard in one or more areas and depths, or could have defined better the small period of high scup discard in area 613. For example, the data indicated that November and December were key months, but more sampling in Octo- ber and January could have widened this time frame. A second problem with the data is that they were collected from many different boats, with different nets, horse powers, tow durations, etc. Although such problems are avoided in fishery-independent surveys by using standard- ized gears and sampling protocols, they are unavoid- able when dealing with observer programs whose objective is to survey normal fishing operations across a variety of vessels, gear-types, etc. in order to de- (SE) n Retained SI 700 - I 600 - ■ Discarded ^ 500 - f\ ra 400 - 1 c 300 - i ^ 1 1 200 - 100 - r c dogfish squid scup skates whiting butterfish red hake fluke monkfish bluefish yellowtall flounder Figure 5 Retained and discard rates (per hour trawled) of the main species caught in those tows done in area 613. between 30 and 40 fm, in the months of November and December ( the identified area and time of consistently high | rates for scup discard). tect fleet-wide patterns. Variation in observer data is inherent in all such programs, and it is only by doing properly designed and replicated, stratified, randomized observer surveys that such problems can be accounted for In most studies that have quantified bycatches, species-specific spatial, and temporal variabilities in 190 Fishery Bulletin 97(1), 1999 discarding often preclude the identification of defi- nite areas and times of persistently high discards of species (see also Robin, 1991; Martinez et al., 1993; Liggins and Kennelly, 1996; Kennelly et al., 1997). The present study shows that high discard rates for scup were rarely consistent in particular areas and times, although one area and time did show some persistence throughout the four years: the relatively small area (area 613) off Long Island, New York, be- tween 30 and 40 fm in the months of November and December each year. It is well known that from May to August each year scup spawn in estuaries, bays, and inshore areas south of Cape Cod (e.g. see Wilk and Brown, 1980; Eklund and Targett, 1990), par- ticularly around Long Island (Finkelstein, 1971). In autumn, after spawning, they migrate south at greater depths towards their wintering grounds from southern New Jersey to Cape Hatteras, returning back to their inshore spawning grounds in spring ( see Neville and Talbot, 1964; Wilk and Brown, 1980; Jeffries and Terceiro, 1985; Eklund and Targett, 1990). It is appar- ent from this behavior, that the occurrence of large numbers of scup as trawl discards in area 613 off Long Island in the autumn of each year coincides with their migration from Long Island's inshore waters to their wintering grounds farther south. When spatial and temporal patterns in observer data reveal persistent areas and times of high dis- card rates, they can be used by fisheries managers to identify where and when various management tools can be applied to reduce discards. These tools usually involve either spatial or temporal closures to fishing (or both) or modifications to fishing gears and practices that reduce discards. If fisheries managers consider the area and time of high scup discard identified in this paper as a can- didate for a closure, the data in Figure 5 provide some information on the consequences this may have on landings of squid and other species. In the identified area and time, an average of over 507 lb of squid were retained per hour towed, whereas an average 319 lb of scup were discarded per hour towed. The actual numbers of fish involved were not available but, because the discarded scup were those individu- als considered too small to retain, discarded weights often represent more individual fish per pound than retained weights. One could expect, therefore, that closing this area at this time would protect signifi- cant numbers of small scup, thereby assisting the 1995 Northeast Regional Stock Assessment Work- shop's recommendation to decrease their exploitation (NEFSCM. Larger quantities of dogfish and lesser quantities of skates, butterfish, and whiting would also be protected by such a closure. The "cost" of such a closure, however, would be significant reductions in the landings of squid from the area at that time and lesser reductions in the landings of scup and whiting. One way of estimating the potential effects that such a closure strategy might have for the re- gion is to compare the overall squid retained and scup discard rates with rates adjusted by excluding all tows done in the identified area and time. Through- out the region, the overall discard rate for scup dur- ing the four years was 37.5 Ib/h (standard error (SE)=4.7), but if the tows done in the identified area and time are excluded, the overall rate falls to 27.3 Ib/h (SE=4.5): a decrease of 27%. This should be com- pared to the overall retained rate of squid in the re- gion of 128.5 Ib/h (SE=6.1) falling to 114.9 Ib/h (SE=6.0) if the tows done in the identified area and time are excluded — a decrease of 10.6%. It is impor- tant to note, however, that closing the area at this time to trawling will not simply remove trawling ef- fort from the region but merely redirect it to other areas that may yield lower scup discards. This means that the effects on landings and discards in the re- gion will not simply be those protected inside the clo- sure but will be tempered by increased landings and discards outside the closure by the redirected vessels. The other suite of management tools available to reduce discards in areas and at times of high dis- carding involve modifying fishing gears and practices to improve selectivity. Such modifications as the Nordmore Grid and square-mesh panels have proven successful in reducing discards of small fish in trawl fisheries (e.g. Carr, 1989; Isaksen et al., 1992; Broadhurst et al., 1996; Broadhurst and Kennelly, 1997), and modifications like downward sorting grids and horizontal panels in nets have reduced the bycatches of unwanted sizes of species in groundfish trawls (see Larsen and Isaksen, 1993; Engas and West"^ ). For the issue of scup discarding in the north- eastern United States, gear modifications would be a better management alternative than the closure strategy outlined above if they could reduce greater quantities of scup discards (and concomitantly re- duce landings of squid and other species by smaller amounts) than those under a closure strategy. Such a general, gear-based solution is also advantageous because its adoption throughout the region would not only lead to reductions in scup discard in identified "hotspots" but in all areas where scup discarding occurs. The information provided in the present study identifies the ideal locations and times for any gear- based research that aims to reduce scup discard by trawlers and where and when any gear modifications ■* Engas, A., and C. W. West. 1995. Development of a species- selective trawl for demersal gadoid fisheries. Int. Coun. Explor Sea Council Meeting (CM) 1995/B+G+H+J+K:1. NOTE Kennelly: Discard rates for Stenotomus chrysops 191 should be initially implemented. At the time of writ- ing this paper, several such gear-based discard reduc- tion programs are currently underway in the region. Acknowledgments This study was funded by Surdna, National Fish and Wildlife, Orchard, and Norcross Foundations. The observer program that provided all data was the National Marine Fisheries Service's Northeast Sea Sampling Program (contract nos.: 50-EANF-8-00089 and 50-EANF-2-00038). I thank Connie Delano- Gagnon for assistance with data handling, Steve Drew, and members of the Sea Sampling Program for helpful discussions, and the many observers in- volved in collecting the data. I also thank Joan Palmer and Daniel Sheehan for providing certain data used, Stacey Grove for producing the map, and Steve Drew, Steve Murawski, Ron Larsen, and three anonymous referees for reviewing the manuscript. Literature cited Alverson, D. L., M. H. Freeberg, S. A. Murawski, and J. G. Pope. 1994. A global assessment of fisheries bycatch and discards. FAO Fish. Tech. Paper 339, 233 p. Anthony, V. C. 1993. The state of groundfish resources off the northeast- ern United States. Fisheries 18(3):12-17. Atkinson, D. B. 1984. Discarding of small redfish in the shrimp fishery ofT Port au Choix, Newfoundland, 1976-80. J. Northwest Atl. Fish. Sci. 5: 99-102. Broadhurst, M. K., and S. J. Kennelly. 1997. The composite square-mesh panel: a modification to codends for reducing unwanted bycatch and increasing catches of prawns throughout the New South Wales oce- anic prawn-trawl fishery. Fish. Bull. 95:653-664. Broadhurst, M. K., S. J. Kennelly, and B. Isaksen. 1996. Assessments of modified codends that reduce the by- catch offish in two estuarine prawn-trawl fisheries in New South Wales, Australia. Fish. Res. 27: 89-112. Cadrin, S. X., A. B. Howe, S. J. Correia, and T . P. Currier. 1995. Evaluating the effects of two coastal mobile gear fish- ing closures on finfish abundance off Cape Cod. N. Am. J. Fish. Manage. 15:300-315. Carr, H. A., ed. 1989. Proceedings of the square mesh workshop. World symposium on fishing vessel and fishing gear design. Department of Fisheries and Oceans, Canada, St. John's, Newfoundland, 133 p. Collins, C. H. 1995. Beyond denial — the northeast fisheries crisis: causes, ramifications, and choices for the future. Fisheries 20( 1 1 ): 4. Eklund, A. M., and T. E. Targett. 1990. Reproductive seasonality of fishes inhabiting hard bottom arteas in the middle Atlantic Bight. Copeia 4: 1180-1184. Finkelstein, S. L. 1971. Migration, rate of exploitation and mortality of scup from the inshore waters of eastern Long Island. New York Fish & Game 18:97-111. Hall, M. A. 1994. Bycatches in purse-seine fisheries. In T. J. Pitcher and R. Chuenpagdee, (eds.), By-catches in fisheries and their impact on the ecosystem, p. 53-58. Fisheries Cen- tre Research Report 2( 1 ), Univ. British Columbia, Canada. Hendrickson, H. M., and W. L. Griffin. 1993. An analysis of management policies for reducing shrimp by-catch in the Gulf of Mexico. N. Am. J. Fish. Manage. 13:686-697. Howell, W. H., and R. Langan. 1987. Commercial trawler discards of four flounder species in the Gulf of Maine. N. Am. J. Fish. Manage. 7:6-17. 1992. Discarding of commercial groundfish species in the Gulf of Maine shrimp fishery. N. Am. J. Fish. Manage. 12:568-580. Isaksen, B., J. W. Valdemarsen, R. B. Larsen, and L. Karlsen. 1992. Reduction of fish by-catch in shrimp trawl using a rigid separator grid in the aft belly. Fish. Res. 13:335- 352. Jean, Y. 1963. Discards offish at sea by northern New Brunswick draggers. J. Fish. Res. Board Can. 20:497-524. Jeffries, H. P., and M. Terceiro. 1985. Cycle of changing abundances in the fishes of the Narragansett Bay area. Mar Ecol. Prog. Ser. 25:239-244. Kennelly, S. J. 1997. A framework for solving by-catch problems: examples from New South Wales, Australia, the eastern Pacific and the northwest Atlantic. In D. A. Hancock, D. C. Smith, A. Grant, and J. P. Beumer (eds.). Developing and sustaining world fisheries resources: the state of science and man- agement, p. 544-550. Proceedings of the 2nd World Fish- eries Congress, Brisbane, Australia. Kennelly, S. J., S. C. Drew, and C. D. Delano Gagnon. 1997. Rates of retained and discarded catches from demer- sal fish trawling off the north-eastern United States. Mar. Freshwater Res. 48:185-199. Larsen, R. B., and B. Isaksen. 1993. Size selectivity of rigid sorting grids in bottom trawls for Atlantic cod {Gadus morhua) and Haddock [Melano- grammus aeglefinus). ICES Mar Sci. Symp. 196:178-182. Liggins, G. W., and S. J. Kennelly. 1996. By-catch from prawn trawling in the Clarence River estuary. New South Wales, Australia. Fish. Res. 25:347- 367. Martinez, E. X., J. M. Nance, Z. P. Zein-Eldin, J. Davis, L. Rathmell, and D. Emiliani. 1993. Trawling bycatch in the Galveston Bay system. Galveston Bay National Estuary Program Publ. GBNEP 34, 185 p. Matthiessen, G. C. 1995. Perspective on finfisheries in southern New England. The Sounds Conservancy Coastal Publication 5, Essex, CT, 48 p. Murawski, S. A., 1992. The challenges of finding solutions in multispecies fisheries. In R. W. Schoning, R. W. Jacobson, D. L. Alverson, T G. Gentle, and J. Auyong, (eds.). Proceedings of the national industry bycatch workshop, February 4-6, 1992, Newport, Oregon, p. 35-45. Natural Resources Consultants Inc., Seattle, WA. 192 Fishery Bulletin 97(1), 1999 1994. Opportunities in bycatch mitigation. In R. H. Stroud, (ed.). Conserving America's fisheries: a national symposium on the Magnuson Act, New Orleans, LA, March 1993, p. 299-311. National Coalition for Marine Conser- vation, Savannah, Georgia. Murawski, S. A., K. Mays, and D. Christensen. 1995. Fishery observer program, /n Status of fishery re- sources off the northeastern United States for 1994, p. 35- 41. U.S. Dep. Commer, NOAA Tech. Memo., National Marine Fisheries Service NE-108. NEFSC (Northeast Fisheries Science Center). 1995. Status of fishery resources off the northeastern United States for 1994. U.S. Dep. Commer, NOAA Tech. Memo. National Marine Fisheries Service NE-108, 140 p. Neville, W. C, and G. B. Talbot. 1964. The fishery for scup with special reference to fluc- tuations in yield and their causes. U.S. Fish Wildl. Serv., Spec. Sci. Rep. Fish. 459, 61 p. Powles, P. M. 1969. Size changes, mortality, and equilibrium yields in an exploited stock of American plaice \Htppoglossoides platessoides). J. Fish. Res. Board Can. 26: 1205-1235. Robertson, J. H. B., and P. A. M. Stewart. 1988. A comparison of size selection of haddock and whit- ing by square and diamond mesh codends. J. Cons. Int. Explor Mer 44:148-161. Robin, J. P. 1991. The brown shrimp fishery of the Loire Estuary: pro- duction and by-catch ofjuvenile fish. Fish. Res. 13:153-172. Shepherd, G. R., and M. Terceiro. 1994. The summer flounder, scup and black sea bass fish- ery of the middle Atlantic Bight and southern New En- gland waters. U.S. Dep. Commer., NOAA Tech. Rep. NMFS 122. 13 p. Sissenwine, M. P., and A. A. Rosenberg. 1994. LIS living marine resources: current status and long- term potential. In R. H. Stroud, (ed.). Conserving America's fisheries: a national symposium on the Magnuson Act, New Orleans, LA, March 1993, p. 29-38. National Coa- lition for Marine Conservation, Savannah, GA. Wilk, S. J., and B. E. Brown. 1980. A description of those fisheries, which take place in the western North Atlantic between the LTS-Canadian bor- der and North Carolina, that presently have or potentially could have user group allocation conflicts. In J. H. Grover, (ed.). Proceedings of the Technical Consultation on Alloca- tion of Fishery Resources, Vichy, France, April, 1980, p. 502-518. FAO, Rome, Young, R. H., and J. M. Romero. 1979. Variability in the yield and composition of by-catch recovered from Gulf of California shrimping vessels. Trop. Sci. 21:249-264. 193 Effects of live-bait shrimp trawling on seagrass beds and fish bycatch in Tampa Bay.. Florida David L. Meyer Mark S. Fonseca Patricia L. Murphey Beaufort Laboratory Southeast Fisheries Science Center National Manne Fisheries Service, NOAA 101 Pivers Island Road Beaufort, North Carolina 28516 E-mail address (for D L Meyer) dave meyenSinoaa gov Robert H. McMichael Jr. Michael M. Byerly Florida Marine Research Institute Florida Department of Natural Resources 100 8'^^ Avenue, S.E. St. Petersburg, Florida 33701 Michael W. LaCroix Paula E. Whitfield Gordon W. Thayer Beaufort Laboratory Southeast Fisheries Science Center National Manne Fishenes Service, NOAA 101 Pivers Island Road Beaufort, North Carolina 28516 The use of live shrimp for bait in recreational fishing has resulted in a controversial fishery for shrimp in Florida. In this fishery, night collec- tions are conducted over seagrass beds with roller beam trawls to cap- ture live shrimp, primarily pink shrimp, Penaeus duorarum . These shrimp are culled from the catch on sorting tables and placed in on- board aerated "live" wells. Beds of turtlegrass, Thalassia testudinum, a species that has highest growth rates and biomass during summer and lowest during the winter (Fon- seca et al., 1996) are predominant areas for live-bait shrimp trawling (Tabb and Kenny, 1969). Because of their use in seagrass beds, roller trawls were designed to roll over the bottom to reduce gear penetration and debris collection. Because turtlegrass has an extensive root system, it is not likely to be up- rooted by roller trawls, but the roller on the trawl has been noted to break off and collect old turtle- grass leaves (Woodbum et al. 1957). On the Gulf Coast of Florida, bait shrimp are generally collected from turtlegrass beds year-round, but most shrimp are taken October through February (Berkeley et al., 1985). Trawl and culling times for this fishery, including that of Tampa Bay, are typically short, 5-20 and 2- 15 min, respectively, to reduce debris collection and both injury and mor- tality to shrimp. Although no data were collected, Berkeley et al. ( 1985) suggested that such trawling for shrimp may be destructive to sea- grass beds and juvenile fishes, in- cluding spotted seatrout, Cynoscion nebulosus, snapper, Lutjanus spp., and pigfish, Orthopristis chrysoptera. However, the effects of this type of trawling on finfish bycatch mortal- ity are unknown. Numerous studies offish bycatch mortality have used otter trawls. These include consideration of the effects otter trawling has on cod, Gadus morhua, and American pla- ice, Hippoglossoides platessoides, in the Gulf of St. Lawrence, Canada (Jean, 1963); on red snapper, Lutjanus campechanus, in the Gulf of Mexico (Gutherz and Pellegrin, 1988); and on bycatch in the Torres Strait, Australia (Wassenberg and Hill, 1989; Harris and Poiner, 1990; Hill and Wassenberg, 1990). In these studies fish mortality ranged from 10% (Jean, 1963) to 80% (Jean, 1963; Wassenberg and Hill, 1989), depending on culling times, animal size, and temperature. Trawl and culling times (30-60 minutes and 15-45 minutes, re- spectively) in these studies were typically longer than those used by the Florida live-bait shrimp fishery. Our study objectives were 1) to determine effects of a roller beam trawl on turtlegrass biomass and morphometries during intensive (up to 18 trawls over a turtlegrass bed), short-term (3-hour duration) use and 2) to examine the mortal- ity of bycatch finfish following cap- ture by a trawl. Methods Sampling was done in August and November 1990 in Tampa Bay, Florida (Fig. 1). A commercial bait shrimp boat towed two 3.38 m wide, 0.8 m high stainless steel roller beam trawls simultaneously, one Manuscript accepted 6 April 1998. Fish. Bull. 97(1):193-199 (1999). 194 Fishery Bulletin 97(1), 1999 from each side of the vessel. These consisted of a net attached to a stainless steel frame with a slotted roller along the entire lower portion of the frame. Stainless steel finger bars, 0.8 m long, were fastened vertically, 5 cm apart, along the front top of the frame to exclude seagrass and other debris (Ber- keley et al., 1985). Each roller beam trawl weighed -75 kg. Nets were constructed of 2.56-cm stretched mesh and had a 1.90-cm stretched mesh tail bag. Turtlegrass Nine trawl areas were selected in turtle- grass beds near Tarpon Key (Fig. 1). Each consisted of a marked 40-m x 3.38-m plot of continuous turtlegrass cover. Within each 40-m plot, a central 20-m x 3.38-m impact area was marked. During August, we sampled turtlegrass by removing cores of it within the impact area of each plot prior to trawling and after 1, 3, and 9 trawls. In November, new plots were marked and samples were collected prior to trawling and again after 9 and 18 trawls. Trawl levels were increased in November because pre- liminary analysis of the August data did not reveal a significant effect on turtlegrass at the maximum trawl level. Five 15-cm diameter x 20-cm deep sub- strate cores were taken randomly along a lengthwise midline transect within each impact area at each specified trawl level. For each core the shoot density, longest blade length, total blade length per shoot, and number of blades per shoot (from three ran- domly selected shoots) were recorded. The turtlegrass standing crop for each core was measured separately for above- and below-ground dry -weight biomass ( dried at 60°C for at least 48 hours). The sequential seagrass measurements within each replicate plot were used in regression analysis to de- tect rates of decrease for each of the different plant parameters as a function of trawl level. The regi-ession slope (rate of decrease) for each plot was calculated for each plant parameter. Mean rate of decrease for each plant parameter was calculated for each month (all plots combined, n=9 for each month). T-tests were used to compare differences (P<0.05 ) in mean rate of decrease for each plant parameter for each month. Mortality of bycatch fish The bycatch collection site was southwest of Tierra Verde, near the mouth of Tampa Bay (Fig. 1). Each bycatch trawl site Figure 1 Location of bycatch and turtlegrass trawl sites within Tampa Bay, Florida, 27 40' North latitude, 82 27' West longitude. month 30 trawls of 5-min duration were made at this site over three consecutive nights. Trawl and culling times selected were based on observations of com- mercial live-bait shrimp fishing vessels and short- est times observed used to obtain conservative esti- mates of bycatch fish mortality. In August, the catch from one net of a trawl pair was allowed to sit on the culling table for 2 min prior to placement in the on- deck live well. The catch from the other trawl net was discarded. While the catch was sitting on the culling table, algae and detritus were removed to reduce subsequent fish entanglement. To reduce the effect of predation on mortality estimates, hardhead catfish, Arius felis, were removed from the catch. Bycatch for each trawl was placed in its own sepa- rate holding pen, which was held on deck inside an aerated 300-L live well until transferred to a final unvegetated holding site. Holding pens along with bycatch contents were transferred from the on-deck live well to the holding site in a 76-L seawater tank. On-deck holding times were 15 min, transport 5 min. Once at the holding site, pens were placed in the water at a low tide depth > 1 m (the minimum) and anchored. NOTE Meyer et al : Effects of live-bait shrimp trawling on seagrass beds and fisfi bycatch 195 Holding pens were cubical, 1.25 m on each side, constructed of 6.4-mm mesh and had removable mesh tops. A 5-cm diameter polyvinylchloride collar was attached around the top sides of the pen to provide floatation. The mesh top and float collar prevented fish from escaping over the sides, and the top pre- vented avian predation. Lead-core line was sewn to the bottom seams of each pen to maintain pen vol- ume. Loops attached to each corner of the pen en- abled us to anchor the pens with conduit poles. Owing to low abundance offish bycatch in Novem- ber, collection methods were modified; the bycatch from both of the paired trawl nets was combined (only a single trawl net was processed in August), and placed into a single holding pen. Handling procedures were also changed to estimate fish bycatch mortal- ity more conservatively by eliminating catch culling time. In November the catch was placed directly into an aerated 76-L transport tank and during trans- port to the hold site, algae, detritus, and hardhead catfish were removed. Thus, bycatch mortality re- sults for the two months were not directly compa- rable. In addition, during November the holding pens were deployed prior to bycatch collection, and the trawl catch was subsequently transferred from the 76-L transport tank into individual holding pens. Pens were checked for dead fish (no opercular movement) following initial placement at the hold site, and then at intervals of 2, 8, 12, 24, and 36 hours. Checks involved inspecting each holding pen by sys- tematically lifting portions of the pen netting so that the entire volume, net sides, and bottom could be visually inspected for dead fish while allowing sur- viving fish to remain immersed. Dead fish were col- lected at each time check and preserved in 10% for- malin. At 36 hours, all dead and live fish were sepa- rated and preserved. Fish were identified, and stan- dard length and body depth were measured (to the nearest 0.5 mm). The number of individuals and weights of each species (to the nearest 0.01 g) were recorded for each trawl and holding pen time check. Fish with a body depth <7.5 mm were excluded from survival measurements because they were able to escape through the mesh and would cause an unrep- resentative mortality estimate. Because the number of individuals per species per trawl was generally <5, we pooled all 30 trawls for each month to assess survival. The more numerous fish species (spotted seatrout, Cynoscion nebulosus, pinfish, Lagodon rhomboides, and silver perch, Bairdiella chrysoura) were subdivided into small (25.0- 65.0 mm standard length, SL) and large (65.5-125.0 mm SL) size classes for mortality comparisons. Mojarra, Eucinostomus spp. (including silver jenny, Eucino- stomus gula), were smaller and were subdivided into small (25.0-55.0 mm) and large (55.5-95.0 mm) size classes so that we could examine size-related mortal- ity. Determination of differential size-class mortality was tested with log-linear model analysis (P<0.05). Salinity and air and water temperatures were measured hourly during bycatch trawling. Results Turtlegrass Regression analysis to estimate mean rate of decrease showed no significant (P>0.05) reduction in mean shoot density, number of blades per shoot, longest blade length per shoot, total blade length per shoot, or above- and below-ground biomass with increased trawling during either month. Monthly vegetation measurements at each trawl level (all nine sites com- bined) are shown in Table 1. Fish bycatch We collected a total of 5901 fish representing 42 spe- cies; 3262 fish (29 species) in August, 2639 fish (36 species) in November. Mojarra (including silver jenny) were 79.3% (August) and 52.1% (November) of the total catch (Table 2). Most mojarra were smaller than 40 mm SL and were identified as Eucinostomus spp., whereas all larger Gerreidae could be identified as silver jenny. The unreliability of the identification of smaller Gerreidae (those smaller than 40 mm SL) precluded their definitive identification to species (Matheson and McEachran, 1984). Mojarra less than 40 mm composed 60.0% (August) and 24.7% (November) of all fish collected, and silver jenny represented 19.3% (August) and 27.4% (November) of all fish collected (Table 2). Survival was variable (0-100% ) among species and months (Table 2; Fig. 2). Among the abundant spe- cies (those with at least 20 individuals for each month), high survival was observed for gulf toadfish, Opsanus beta, pigfish, pinfish, and gray snapper, (Table 2; Fig. 2). Abundant species with low survival were silver perch, mojarra, silver jenny, and spotted seatrout. Greatest mortality in most species occurred within the first 8 or 12 hours after collection (Fig. 2). Small fish were significantly (P<0.05) more sus- ceptible to trawl-induced mortality than larger fish (Table 3). This was particularly evident for silver perch, spotted seatrout, pinfish, and mojarra (includ- ing silver jenny). Species-specific survival was greater in November than August (Table 2; Fig. 2). For those species for which at least 20 individuals were collected each 196 Fishery Bulletin 97(1), 1999 month, an average increase of 30.5% species-specific survival was observed in November compared with August. Silver perch (63.6%), gulf pipefish, Syng- nathus scovelli, (48.1%), and gray snapper (35.0%) had the highest survival increases. Average air and water temperatures observed were 31 and 32°C, respectively, in August, 23 and 24°C, respectively, in November. Salinity averaged 35 ppt during both months. Discussion Because roller trawls were designed to reduce seagrass fragment collection, it has been assumed that they have minimal impact on seagrass habitat (Tabb and Kenny, 1969; Berkeley et al., 1985). We were unable to detect significant trawl impacts on shoot den- sity, structure, or biomass of turtlegrass by intensive short-term ( 18 trawls within three hours ) trawling. This finding supports conjectures by Woodbum et al. ( 1957) and Tabb and Kenny (1969) that roller trawls have minimal impact on turtlegrass habitat. However, we did not test for effects of repetitive trawling over turtlegrass beds over a longer period of time. Trawling may cause elevated localized tur- bidity, which could chronically reduce the potential for seagrass to perform photosynthesis (Kenworthy and Haunert, 1991). In contrast, limited trawling in areas with substantial epiphytic growth and numer- ous senescent turtlegrass leaves may enhance light availability by removing old plant parts (Woodbum 1» 90 80 70 40 3& zy vy a lOOn 90 80 -•- -■A- 70- 4 60 50- 40- gut X BUP 20 10- n- =»= 05 2A ao IZA 24.0 36.0 2.0 8.0 24i) 36J) E u O lOOi 90 80 -»- an A- 70 spo 60 --»-- 50 -B- 40 gra 30 pig 20 10- 0 November D aO MB »0 36.0 0.5 12.0 Hours after trawling Figure 2 lA-D) Cumulative percent mortality in August and November of the abundant (at least 20 individuals per sample month I species collected with shrimp trawls and held in floating pens for 36 h. As noted in the text, handling procedures differed between the two collection periods. Species codes: moj = mojarra, sij = silver jenny, gol = goldspotted killifish, gut = gulf toadfish, gup = gulf pipefish, sil = silver perch, spo = spotted seatrout, pin = pinfish. gra = gray snapper, and pig = pigfish. NOTE Meyer et al.: Effects of live-bait sfirimp trawling on seagrass beds and fish bycatcfi 197 et al., 1957) and epiphytic growth from the plants (author's personal observation). Trawling may affect associated faunal communi- ties by collecting and redistributing macroalgae and turtlegrass litter. Redistribution can reduce the habi- tat complexity of one area and increase that of an- other. The alteration of habitat can in turn influence species composition and abundance (Gore et al., 1981; Kulczycki et al., 1981; Leber, 1985). Fish species were not equally susceptible to mor- tality by exposure or net injury. Regardless of han- dling procedures, mojarra were highly susceptible to mortality. At the other extreme, oyster toadfish and striped burrfish, Chilomycterus schoepfi, were much less susceptible to trawl-induced mortality. Because of differential species mortality, fish species diver- sity and composition may be altered in areas of in- tense or long-term trawling. Mortality and size were inversely related for nu- merous species, also noted by Jean ( 1963) for cod and American plaice, and by Fritz and Johnson (1987) for freshwater drum, Aplodinotus grunniens. A pos- Table 1 Mean (±1 SE) shoot density, morphometries, and biomass of turtlegra is per 176.7 cm- at selected trawl levels during August and November. For each trawl level d jring a month n = 9. Number of Longest Total length Number of Shoot blades blade per all blades Above-ground Below-ground trawls density per shoot shoot (cm) per shoot (cm) biomass (g) biomass (g) August 0 8.1(0.8) 2.5(0.1) 31.6(3.0) 63.5(7.4) 1.3(0.1) 7.4(0.6) 1 7.1(1.1) 2.4(0.1) 32.4(2.0) 61.3(3.7) 2.3(0.4) 7.0(1.1) 3 7.8(0.9) 2.4(0.1) 34.0(2.6) 64.4(6.4) 2.2(0.2) 7.2(1.0) 9 9.1 (0.9) 2.4(0.1) 28.3(1.6) 55.2(3.3) 1.5(0.1) 8.3(0.8) November 0 10.2(0.7) 2.3(0.1) 24.1(1.2) 41.4(2.7) 1.1(0.1) 7.1(0.6) 9 10.4(0.41 2.3(0.1) 23.7(0.9) 40.2(2.0) 1.2(0.1) 7.5(0.5) 18 10.4(0.6) 2.2(0.1) 23.3(1.4) 38.3(2.3) 1.1(0.1) 7.3(0.6) Table 2 Fish collected during Augu st and November live-bait shrimp traw ling in Tampa Bay. Florida . Percent survival was determined 36 hours after collection. Total number Percent of Percent Common name Scientific name ca ught total catch survival August November August November August November Mojarra (<40 mm) Eucinostomus spp. 1956 655 60.0 24.7 0.5 18.2 Silver jenny Eucinostomus gula 629 725 19.3 27.4 3.2 45.7 Goldspotted killifish Floridichthys carpio 182 82 5.6 3.1 36.8 65.0 Pmfish Lagodon rhomboides 152 259 4.7 9.8 71.7 98.1 Scaled sardine Harengula jaguana 52 11 1.6 0.4 15.8 54.6 Silver perch Bairdiella chrysoura 46 77 1.4 2.9 0.0 63.6 Rainwater killifish Lucania parva 42 9 1.3 0.3 28.6 44.4 Gulf toadfish Opsanus beta 40 101 1.2 3.8 97.5 98.0 Pigfish Orthopristis chrysoptera 32 122 1.0 4.6 84.4 95.9 Spotted seatrout Cynoscion nebulosus 25 117 0.8 4.4 28.0 59.0 Gulf pipefish Syngnathus scovelli 22 33 0.7 1.2 45.8 93.9 Gray snapper Lutjanus griseus 20 46 0.6 1.7 65.0 100.0 Striped burrfish Chilomycterus schoepfi 13 198 0.4 7.5 100.0 99.5 Blackcheek tonguefish Symphurus plagiusa 11 21 0.3 0.8 63.6 90.5 Planehead filefish Monacanthus hispidus 1 84 <0.1 3.2 0.0 95.2 Southern puffer Sphoeroides nephelus 0 26 0.0 1.0 96.2 Others 39 73 1.2 2.8 33.3 82.2 198 Fishery Bulletin 97(1), 1999 sible explanation for perceived differences in trawl- induced mortality between small and large size-class fish may be attributed to larger fish having a greater proportion of muscle tissue and larger energy stores than smaller fish; the larger fish are thus able to avoid net contact during prolonged trawl tows, (Fritz and Johnson, 1987). This may cause small fish to be more susceptible than large fish to stress and mor- tality due to capture. Consequently, high mortality of small individuals, directly or indirectly due to trawling, could potentially reduce the local stock of juvenile fish. This reduction in juvenile fish in turn could reduce the reproductive potential of a species and alter species diversity and composition within affected areas (Wassenberg and Hill, 1989). Greater survival during November was most evi- dent for silver perch, gulf pipefish, and gray snap- per. However, we were not able to assess whether observed increased survival was the result of changes in handling procedures that we instituted (culling versus no culling), or to seasonal factors, such as dif- ferences in water temperature as noted by Jean (1963), between our survey months. The 2-min cull- ing time and high August air temperature may have substantially stressed the fish in our study. Also, the Table 3 Percent fish mortality in relation to standard length (mm ) for species with 20 or more individuals collected during each month. Parentheses indicate number of individuals in a size class. Asterisks ( * ) indicate significant differences | P<0.05), based on log-linear model analysis, in mortality between size classes. Size class (mm) Collection Species period 25.0-65.0 65.5-125.0 Silver perch August 100.0 100.0 (42) (4) November* 46.0 14.8 (50) (27) Spotted seatrout August* 65.0 20.0 (20) (5) November* 52.6 17.9 (78) (39) Pinfish August* 47.1 22.9 (34) (118) November 1.5 Mojarra August (259) 25.0-55.0 55.5-95.0 99.1 92.2 ■(2435) (153) November* 71.0 28.0 (12.56) (125) higher water temperature in August may have re- duced the ability offish to recuperate from handling because of increased respiratory demand ( Alderdice, 1963; Bond, 1979) and lower ambient dissolved oxy- gen levels (Raymont, 1980). However, even with the changes in handling instituted in November to re- duce stress, net injury alone appeared to cause sub- stantial mortality in mojarra, spotted seatrout, sil- ver perch, and goldspotted killifish, Floridichthys carpio. Mean size of fish caught was larger in No- vember than in August; this too may have contrib- uted to the higher survival observed. Fish abundance per holding pen was not recorded until the end of the 36-h mortality observation pe- riod. Because of this, and the fact that predation of live and dead individuals within the pens was not measured, the mortality for smaller species or smaller individuals of a species may be conservative. However, the initial confinement may have also en- hanced the survival of injured or disoriented fish by reducing predation that might have been experienced in an open system. Even if bycatch organisms sur- vive trawling and culling they may be quite suscep- tible to predation. High postrelease predation on bycatch has been observed by prawn trawlers (Hill and Wassenberg, 1990), and this fate probably awaits much of the bycatch overboarded during bait-shrimp trawling. Many bait-shrimp boat captains have re- ported schools of hardhead catfish following boats and feeding on the bycatch as it is thrown overboard. We similarly observed numerous hardhead catfish and other predatory fish following our boats and feed- ing on bycatch, and numerous such predatory indi- viduals were collected during our sampling. Trawl-induced mortality may occur past our 36-h observation period. Numerous fish still alive after the 36-h observation period had missing caudal fins or ulcers, or both. Such fish are were susceptible to subsequent mortality through predation or infection, or both. Moreover, individuals recaptured and culled again in heavily exploited areas could suffer high mortality rates because of cumulative stress. To un- derstand better the effects of this fishery on the resi- dent fauna, we need additional evaluation of the ef- fects of temperature on mortality; the consequences of postrelease predation, delayed mortality due to initial sublethal damage, and subsequent infection; and finally an evaluation of the potential for recap- ture of bycatch fish. Acknowledgments We thank D. Colby, M. Hall, M. Durako, and S. Kennedy for advice and guidance on the study de- NOTE Meyer et al.: Effects of live-balt shinmp trawling on seagrass beds and fish bycatch 199 sign and implementation, S. Peck for field assistance, and C. Lewis for assistance with graphics. This study would not have been possible without the coopera- tion and assistance of numerous live-bait-shrimp boat captains who took an interest in this study. We thank W. Hettler, W. Kenworthy, M. Johnson, and three anonymous reviewers for editorial comments. The project was supported through Florida Department of Natural Resources contract number C4488 and the National Marine Fisheries Service. Literature cited Alderdice, D. F. 1963. Some effects of simultaneous variation in salinity, temperature and dissolved oxygen on the resistance of young coho salmon to a toxic substance. J. Fish. Res. Board Can. 20:525-.550. Berkeley, S. A., D. W. Pybas, and W. L. Campos. 1985. Live bait shrimp fishery of Biscayne Bay. Florida Sea Grant Ext. Prog. Tech. Pap. 40. 16 p. Bond, C. E. 1979. Biology of fishes. Saunders Inc., Philadelphia, PA, 514 p, Fonseca, M. S., W. J. Kenworthy, and F. X. Courtney. 1996. Development of planted seagrass beds in Tampa Bay, Florida, USA. I. Plant components. Mar Ecol. Prog. Sen 132:127-139. Fritz, K. R., and D. L. Johnson. 1987. Survival of freshwater drums released from Lake Erie commercial shore seines. N. Am. J. Fish. Manage. 7:293- 298. Gore, R. H., E. E. Gallaher, L. E. Scotto, and K. A. Wilson. 1981. Studies on decapod Crustacea from the Indian River Region of Florida; XI. Community composition, structure. biomass and species-areal relationships of seagrass and drift algae-associated macrocrustaceans. Estuarine Coastal Shelf Sci. 12:48,5-508. Gutherz, E. J., and G. J. Pellegrin. 1988. Estimate of the catch of red snapper, Lutjanus campechanus. by shrimp trawlers in the U. S. Gulf of Mexico. Fish. Rev. 50(l):17-25. Harris, A. N., and L R. Poiner. 1990. Bycatch of the prawn fishery of Torres Strait; composi- tion and partitioning of the discards into components that float or sink. Aust. J. Mar Freshwater Res. 41:37-52. Hill, B. J., and T. J. Wassenberg. 1990. Fate of discards from prawn trawlers in Torres Strait. Aust. J. Mar Freshwater Res. 41:53-64. Jean, Y. 1963. Discards offish at sea by northern New Brunswick draggers. J. Fish. Res. Board Can. 20:497-525. Kenworthy, W. J., and D. Haunert (eds.). 1991. The light requirements of seagrasses: proceedings of a workshop to examine the capability of water quality cri- teria, standards and monitoring programs to protect seagrasses. LIS. Dep. Commer. NOAA Tech. Memo. NMFS-SEFC-287, 181 p. Kulczycki, G. R., R. W. Virnstein, and W. G. Nelson. 1981. The relationship between fish abundance and algal biomass in a seagrass-drift algae community. Estuarine Coast. Shelf Sci. 12:341-347. Leber, K. M. 1985. The influence of predatory decapods, refuge, and microhabitat selection on seagrass communities. Ecology 66:1951-1964. Matheson, R. E., and J. D. MeEachran. 1984. Taxonomic studies of the Eucinostomus argenteus complex (Pisces; Gerreidae); preliminary studies of the external morphology. Copeia 4:893-902. Raymont, J. E. 1980. Plankton and productivity in the oceans. 2nd ed., vol. 1, Phytoplankton. Pergamon. Oxford, 489 p. Tabb, D. C, and N. Kenny. 1969. A brief history of Florida's live bait shrimp fishery with description of fishing gear and methods. FAO Fish Rep. 57:1119-1134. Wassenberg, T. J., and B. J. Hill. 1989. The effects of trawling and subsequent handling on the survival rates of the bycatch of prawn trawlers in Moreton Bay. Australia. Fish. Res. 7:99-110. Woodburn, K. D., B. Eddred, E. Clark, R. F. Hutton, and R. M. Ingle. 1957. The live bait shrimp industry of the west coast of Florida. Florida State Board Conserv. Tech. Ser 21, 33 p. 200 Distribution and relative abundance of sea turtles caught incidentally by the U.S. pelagic longline fleet in the western North Atlantic Ocean, 1992-1995 Wayne N. Witzell Southeast Fisheries Science Center National Marine Fisheries Service, NOAA 75 Virginia Beach Drive Miami, Florida 33149-1099 E-mail address wayne witzellm'noaa gov aboard selected longline vessels (Lee et al., 1995; Gerrior, 1996). Sea turtle species identifications from the logbook data were edited to in- clude only leatherback, Dermochelys coriacea, and loggerhead, Caretta caretta, sea turtles. This decision was based on the known distribu- tion, abundance, and biology of sea turtles in the area, and the fact that some vessel captains and NMFS observers were unable to identify accurately all turtles encountered. The distribution and abundance of threatened and endangered species of sea turtles in offshore waters are not well understood. Early oceano- graphic flights designed to record sea surface temperatures along the Atlantic continental shelf by the U.S. Coast Guard illustrated that aerial surveys for large pelagic fish, marine mammals, and sea turtles were possible ( Deaver' ). Aerial sur- veys have since proven a cost-effec- tive method of obtaining observa- tional data on sea turtles and have helped researchers understand basic distributional patterns (Hoff- man and Fritts, 1982; Fritts et al.. 1983; Schroeder and Thompson, 1987; Shoop and Kenney, 19921. Additional sources of pelagic sea turtle data are fishery observer and vessel logbook programs. The U.S. Atlantic pelagic longline fishery for tuna, Thunnus spp., and swordfish, Xiphias gladius, incidentally cap- tures threatened and endangered sea turtles, which have either in- gested baited hooks or become en- tangled or hooked externally, or both. This paper examines the sea- sonal distribution and relative ' Deaver, J. W. 1975. Aerial oceano- graphic nb.servation.s, July 1969-.Junc 1970, Cape Cod, Massachusetts to Miami, Florida. U.S. Dep. Trans., Coast Guard Oceanogr. Unit, Oceanogr. Rep. CG .373- 68, 27 p. INTIS Accession No. AD-A014 415.1 abundance of these turtles caught incidentally by the U.S. Atlantic pelagic longline fleet from 1992 through 1995. Materials and methods The fishery data used in this analy- sis are from the National Marine Fisheries Service (NMFS) pelagic logbook program managed by the Southeast Fisheries Science Cen- ter, Miami Laboratory. The logbook program was initiated in 1991 and requires U.S. Atlantic longline ves- sels to report daily catch and effort data. Specific longline information from the pelagic logbooks includes target species, type of bait, set and haulback dates and positions, length of mainline, number and lengths of gangions and floatlines, and numbers of light sticks and hooks set (Cramer, 1996). Sea turtle bycatch information was added to the logbook data form in 1992. Only sets targeting tuna or swordfish, or both, are analyzed here because the shark longline fishery is signifi- cantly different temporally, spa- tially, and geographically. NMFS logbook data for 1992-95 were ana- lyzed by nine geographic fishing areas established by the NMFS Miami Laboratory (Fig. 1). Biologi- cal data of captured turtles were pro- vided by NMFS observers placed Results and discussion The typical Lf.S. pelagic tuna and swordfish longline consists of a mainline, suspended horizontally by floats at a set depth, that has a series of baited hooks hanging ver- tically. The gears used for both tar- get species are essentially the same. However, swordfish fisher- men generally work at night using chemical light sticks suspended above the hooks to attract bait fish and tuna fishermen generally fish during the day without light sticks. Still, there are indications from log- book data that some of the longline fleet apparently target both tunas and swordfish and alter fishing strategies and gear configurations depending on the main target spe- cies, geographic location, and sea- son (Hoey, 1983; Sakagawa et al., 1987). The U.S. pelagic longline fishery has evolved over the years from cumbersome New England style gears to lighter Florida style gears (Berkeley etal., 1981)andare significantly different from the gears used by the Japanese who targeted bluefin tuna in the west- ern Atlantic (Lopez et al., 1979; Witzell, 1984). The gears currently in use on U.S. vessels vary consid- erably; they average about 47 km long and have 429 hooks suspended by 174 floats 54 m below the sur- Manuscript accepted 24 April 1998. Fish. Bull. 97:200-211 (19991. NOTE Witzell: Distribution and abundance of sea turtles caught in the western Atlantic Ocean 201 1 - Caribbean 2 - Gulf of Mexico 3 - Florida east coast 4 - South Atlantic bight 5 - Mid-Atlantic bight 6 - North east coastal 7 - North east distant 8 - North equatorial 9 - Mid-Atlantic Ocean Figure 1 The NMFS pelagic fishing areas. Dashed line represents the 200-m isobath. face (Power-). Pelagic longline fishing effort also changes temporally and spatially, depending on tar- get species, season, and location (Fig. 2), although the overall effort for the entire area has not changed significantly over recent years (Cramer, 1996). Most effort is closely associated near the 200-m isobath along the edge of the continental shelf off the U.S. and off the shelf near the southeastern edge of the Grand Banks and often coincides with major current systems and thermally dynamic areas (Podesta et al., 1993). The most noticeable seasonal shift in fishing ef- fort is the summer-fall swordfish fishery at the Grand Banks that moves to the Caribbean in the winter. Leatherback sea turtles A total of 1264 leatherback sea turtle captures were recorded in the NMFS pelagic logbooks for 1992-95 (Table 1). Total numbers of individual turtles caught ^ Power, J. H. 1995. Analysis of the longline fishery effort, catch, and bycatch in the southwest Atlantic and Gulf of Mexico. U.S. Dep. Commer. Natl. Mar Fish. Serv. Southeast Regional Office, 9721 Executive Center Drive. North, St. Petersburg, FL 33702. NOAA-NMFS-MARFIN Final Report, NA37FF0040- 01, 96 p. are unknown because some turtles may have become entangled more than once. Year-round leatherback captures averaged 316 individuals per year (range: 198-452); June through November were the most important months with 1047 captures (82.7%). The mid-Atlantic, northeast coast, and northeast distant waters (areas 5-7) were the most productive, with 1035 (81.9%) combined captures. Of these, the north- east distant (area 7) accounted for 593 (46.8%) of the total leatherback turtle captures. Leatherback sea turtles are captured sporadically throughout the fish- ing area in the winter and spring seasons and be- come more abundant during the summer and fall, particularly in fishing areas 5-7 (Fig. 3). September was the most productive month with 278 (21.9%) captures for these areas combined. Reported catches varied annually (Table 2). Leatherback sea turtles were too large for observers to bring on deck for length and weight measurements, but the mean es- timated carapace length of 1 10 individuals observed from the decks of northwest Atlantic fishing vessels was 160 cm. Catch-per-unit-of-effort (CPUE) values indicated that loggerhead sea turtle capture rates varied con- siderably between areas (Table 3) but were highest 202 Fishery Bulletin 97(1), 1999 Effort (longline sets) Spring 1992 - 95 K** Effort (longline sets) Summer 1992 - 95 Figure 2 Pelagic longline fishing effort (sets), by seiiscm. Dashed line represents the 200-ni isobath. NOTE Witzell: Distribution and abundance of sea turtles caught in the western Atlantic Ocean 203 Effort (longline sets) Fall 1992 - 95 ^'^r ^ Effort (longline sets) Winter 1992 - 95 ^1:. :^$f^ Figure 2 (continued) 204 Fishery Bulletin 97(1), 1999 Leatherback Spring 1 992 - 95 ^' -sz^-''^'- -??" >-.. -^1. r" ^' Leatherback Summer 1992 - 95 v/ t ■■- .-sz^'. ~sv^. 5**^ •^- ■ '<)*:^:^ . Figure 3 Leatherback sea turtle catches, by season (1992-95). Dashed line represents the 200-m isobath. NOTE Witzell: Distribution and abundance of sea turtles caught in the western Atlantic Ocean 205 Leatherback Fall 1992 - 95 V 1/ 0 ^C>-^"-. "X ■ / :^ ■■■-.. ^ Leatherback Winter 1992 - 95 ^": -^- 'lt> Figure 3 (continued) 206 Fishery Bulletin 97(1), 1999 Table 1 Leatherback (Lb) and lo ggerhead (Lh) turtle captures by the U.S. pelagic longl ine fleet, by NMFS fishing area ■ by month, for | 1992-95 All areas Month Area 1 Area 2 Area 3 Area 4 Area 5 Area 6 Area 7 Area 8 Area 9 combined Lb Lh Lb Lh Lb Lh Lb Lh Lb Lh Lb Lh Lb Lh Lb Lh Lb Lh Lb Lh 01 16 12 4 6 4 2 0 0 1 4 0 0 0 0 0 0 5 4 30 28 02 17 7 12 1 2 3 0 0 0 7 0 0 0 0 0 0 2 3 33 21 03 13 24 8 4 6 2 3 1 6 7 0 0 0 0 4 2 0 1 40 41 04 4 4 7 3 2 1 3 2 1 3 2 1 0 0 1 1 0 0 20 15 05 5 1 1 4 1 1 7 4 13 3 5 3 0 0 11 1 0 0 53 17 06 1 1 3 9 2 1 6 3 22 24 35 49 40 39 3 0 1 0 113 126 07 2 0 10 5 0 0 3 3 19 10 48 62 170 233 0 0 0 0 252 313 08 3 0 10 7 1 0 2 2 32 24 48 14 152 212 0 0 0 0 248 259 09 0 1 0 1 1 1 4 2 76 11 30 4 167 310 0 0 0 0 278 330 10 1 1 2 5 1 0 0 3 84 8 12 6 56 123 0 0 0 0 156 146 11 3 1 3 2 0 1 2 0 7 5 0 4 8 19 0 0 2 0 25 32 12 4 4 3 1 3 1 0 0 1 2 0 0 0 0 0 0 5 1 16 9 Total 69 56 73 48 23 13 30 20 252 108 180 143 593 936 19 4 15 9 1264 1337 in the northeast distant (area 7). The overall CPUE for leatherback sea turtles caught by longline ves- sels using chemical light sticks was higher than the CPUE with vessels not using light sticks (Table 3), suggesting that the chemical light sticks might simu- late bioluminescent gelatinous prey and attract leatherback sea turtles to the branchlines. Fortu- nately, these turtles tend to become entangled in the branchlines rather than consume baited hooks. Leatherback sea turtles inhabit coastal and off- shore pelagic waters in the North Atlantic Ocean (Pritchard, 1971). Morreale et al. (1994) speculated that they migrated along specific bathymetric con- tours outside the 200-m isobath, and Lutcavage (1996) proposed that they seek high concentrations of gelatinous prey along the oceanic fronts and me- anders of the Gulf Stream. Shoopand Kenney (1992) reported that most were scattered along the conti- nental shelf, except off Rhode Island in 1978, when a number of individuals were observed in associa- tion with a large concentration of Cyanea. Pelagic longline data, however, suggest that large numbers of leatherback sea turtles apparently also inhabit deep waters extending over the edge of the continen- tal shelf outside the 200-m isobath from Cape Hatteras, North Carolina, to Georges Bank and the Grand Banks in the summer and fall months. Loggerhead sea turtles A total of 1337 loggerhead sea turtle captures were re- corded in NMFS pelagic logbooks, 1992-95 (Table 1). Table 2 Leatherback (Lb) and loggerhead (Lh) turtle captures by | the U.S pelagic long ine fleet, by NMFS fishing area, by year. Area 1992 1993 1994 1995 Lb Lh Lb Lh Lb Lh Lb Lh 1 19 12 17 6 8 11 25 27 2 20 9 25 8 12 6 16 25 3 3 2 6 4 11 3 3 4 4 10 2 3 6 7 5 10 7 5 147 30 68 13 29 32 18 33 6 81 26 54 51 19 19 26 47 7 84 59 67 33 105 278 337 566 8 0 0 0 1 3 0 16 3 9 5 0 5 5 4 2 1 2 Total 369 140 245 127 198 356 452 714 Total numbers of individual turtles caught are un- known because observer data indicate that some turtles may have been captured more than once. Year- round loggerhead sea turtle captures averaged 334 individuals per year (range: 127-714); June through November was the most productive period with 1174 (87.8%) captures. The mid-Atlantic Bight, northeast coast, and northeast distant waters (areas 5-7) were the most productive with a combined total of 1187 (88.7%) loggerhead sea turtle captures. Of these, the northeast distant area (7) accounted for 936 (70.0%) of the captures. Like leatherback sea turtles, logger- NOTE Witzell: Distribution and abundance of sea turtles caught in the western Atlantic Ocean 207 Table 3 Leatherback (Lb) and loggerhead (Lh) turtle ncidental CPUE from 1992- -95 pelagic logbooks by area, with (Y) and without (N) chemical light sticks. CPUE is turtle capt ures per 1000 hooks fished. Lb Lh Lb Lh Area Lights Hooks Captures Captures CPUE CPUE 1 Y 1,036,600 66 54 0.0637 0.0521 1 N 101,716 3 2 0.0295 0.0197 2 Y 2,580,340 38 37 0.0147 0.0143 2 N 3,132,180 35 11 0.0112 0.0035 3 Y 1,389,240 23 13 0.0166 0.0094 3 N 45,773 0 0 0.0000 0.0000 4 Y 937,620 30 19 0.0320 0.0203 4 N 88,089 0 0 0.0000 0.0114 5 Y 1,523,970 154 53 0.1011 0.348 5 N 2,052,910 108 55 0.0526 0.0268 6 Y 1.356,540 132 83 0.0973 0.0612 6 N 835,484 48 60 0.0575 0.0718 7 Y 2,086,660 591 932 0.2832 0.4466 7 N 55,214 2 4 0.0362 0.0724 8 Y 51,593 18 4 0.3489 0.0774 8 N 1,285 1 0 0.7782 0.0000 9 Y 497,265 15 9 0.0302 0.0181 9 N 25,696 0 0 0.0000 0.0000 1-9 Y 11,459,800 1067 1204 0.0931 0.1051 1-9 N 6,338,350 197 133 0.0311 0.0210 head sea turtles were sporadically captured through- out the entire fishing area in the winter and spring and became abundant during the summer and fall in fishing areas 5-7 (Fig. 4). September was the most productive month with a combined total of 330 (24.7%) loggerhead sea turtle catches. Reported catches varied annually (Table 2). The mean curved carapace length (nuchal notch to tip of marginal) and weights of these captured sea turtles from the north east distant area (7) were 55.9 cm (SD=6.5 cm, n=98) and 22.1 kg (SD=11.1 kg, /)=72). These turtles were smaller than coastal loggerheads from the northeast- ern United States (Lutcavage and Musick, 1985). CPUE values indicated that loggerhead capture rates varied between areas (Table 3) and were high- est for the northeast distant area (7). CPUE analy- sis indicated that the overall loggerhead sea turtle capture rates with light sticks were higher than the CPUE without light sticks (Table 3). Unlike leather- back sea turtles, loggerhead sea turtles readily con- sumed baited hooks. The fate of such hooked turtles is unknown, but they have been observed trailing multiple gangions; therefore some turtles previously hooked have remained actively feeding in the area. Pelagic longline data indicate that loggerhead turtle distribution extends from coastal waters to the edge of the continental shelf from Cape Hatteras, North Carolina, to Georges Bank and the Grand Banks in the summer and fall months and also ex- tends into deep waters off the shelf These turtles are apparently able to move out of critically cold win- ter slope water to more hospitable southern waters. NMFS tagging data revealed that turtles hooked on their flippers were tagged in the spring in areas 6 and 7 and recovered in warmer winter waters in ar- eas 4 and 2, respectively (NMFS'^). Conclusions It is not surprising that pelagic longline fisheries incidentally catch turtles because fishing effort is concentrated either near the 200-m isobath or near interactions between major current systems. In these areas a variety of potential prey species (various nek- tonic organisms associated with Sargassum, as well as cephalopods, coelenterates, and schooling fish) concentrate and consequently attract tunas, sword- fish, and sharks (Laurs et al., 1984; Maul et al., 1984; Kurowicki, 1987; Fiedler and Bernard, 1987; Podesta et al., 1993; Lutcavage, 1996). Aerial survey data off the eastern United States (Shoop and Kenney, 1992) and pelagic longline data presented here indicate that ■^ NMFS Cooperative Marine Turtle Tagging Program, Miami Laboratory, 75 Virginia Beach Drive. Miami, PL 33149. 208 Fishery Bulletin 97(1), 1999 Loggerhead Spring 1992 - 95 Loggerhead Summer 1992 - 95 Figure 4 Loggerhead sea turtle catches, by season 1 1992-95). Dashed line represents the 200-m isobath. NOTE Witzell: Distribution and abundance of sea turtles caught in the western Atlantic Ocean 209 Loggerhead Fall 1992 - 95 Loggerhead Winter 1992 - 95 Figure 4 (continued) 210 Fishery Bulletin 97(1), 1999 leatherback and loggerhead sea turtles use the en- tire continental shelf, as well as the thermally dy- namic waters off the continental edge. The northeast distant fishing area (7) is particu- larly productive. The circulation and thermodynam- ics of fishing area 7 are very complex and provide a unique pelagic habitat. The Gulf Stream begins to meander northeastward off Cape Hatteras, and the western edge of the Gulf Stream eventually passes south of the Grand Banks as it turns eastward to- ward Europe. The thermodynamics of north Atlan- tic circulatory process near the Grand Banks is com- plicated (Schmitz and McCartney, 1993), and it is this interaction of cold slope water and warm Gulf Stream water that is productive for large pelagic fish and sea turtles. Moreover, the Gulf Stream spins off warm-core rings on the cooler northeastern slope water (Auer, 1987), where pelagic longline fishing is heaviest. These warm-core rings have been well stud- ied, resulting in a plethora of published literature (Wiebe and McDougall, 1986). Unfortunately, these biological studies have concentrated on planktonic productivity and mesopelagic fishes rather than on larger pelagic apex predators; however, these rings affect the distribution and abundance of large pe- lagic fish, marine mammals, and sea turtles. Fishery managers and sea turtle researchers need to develop conservation strategies that mitigate sea turtle and longline interactions. The high numbers of potentially lethal loggerhead sea turtle captures in the northeast distant area (7) during the summer and fall swordfish fishery particularly need to be addressed by fishery managers. Acknowledgments I thank J. Cramer, for providing turtle data and back- ground information on the U.S. longline fleet and the NMFS pelagic logbook database, and H. Huang, for providing the tables and graphics. The manuscript was improved by comments fi-om D. Lee, M. Lutcavage, W. Richards, and two anonymous reviewers. Literature cited Auer, S. J. 1987. Five year climatological survey of the gulf .stream sy.stem and it.s as.soc!ated rings. J. Geophys. Res. 92:11,709-11.726. Berkeley, S. A., E. W. Irby, and J. W . JoUey. 1981. Florida's commercial swordfish fishery: longline gear and methods. Florida Cooperative Extension Ser\'ice, Univ. Miami, Miami, FL, Florida Sea Grant Marine Advi- sory Bull. MAP-14, 23 p. Cramer, J. 1996. Large pelagic logbook newsletter-1995. U.S. Dep. Commer, NOAA Tech. Memo. NMFS-SEFSC-394. 28 p. Fiedler, P. C, and H. J. Bernard. 1987. Tuna aggregation and feeding near fronts observed in satellite imagery Continental Shelf Res. 7:871-881. Fritts, T. H., W. Hoffman, and M. A. McGehee. 1983. The distribution and abundance of marine turtles in the Gulf of Mexico and nearby Atlantic waters. J. Herpetol. 17:327-344. Gerrior, P. 1996. Incidental take of sea turtles in northeast U.S. waters. /;; P. Williams, P. A. Anninos, P. T. Plotkin, and K. L. Salvini (compilers). Pelagic longline fishery interac- tions., p. 14-31. U.S. Dep. Commer., NOAA Tech. Memo. NMFS-OPR-7. Hoey, J. 1983. Analysis of longline fishing effort for apex predators (swordfish, shark, and tuna) in the western North Atlan- tic and Gulf of Mexico. Ph.D. diss., Univ. Rhode Island. Kingston, RI, 288 p. Hoffman, W., and T. H. Fritts. 1982. Sea turtle distribution along the boundary of the Gulf Stream current off eastern Florida. Herpetol. 38:40.5-409. Kurowicki, A. 1987. Effect of environmental conditions on concentration formation in tunas and sharks in the central Atlantic ( 1981-198.5). Bull. Sea Fish. Inst. 18:37-43. Laurs, R. M., P. C. Fiedler, and D. R. Montgomery. 1984. Albacore tuna catch distributions relative to environ- mental features observed from satellites. Deep-Sea Res. 31:1088-1099, Lee, D. W., C. J. Brown, and T. L. Jordan. 1995. SEFSC pelagic observer program data summary for 1992-1994. U.S. Dep. Commer. NOAA Tech, Memo, NMFS-SEFSC-373. 19 p. Lopez, A. M., D. B. McClellan, A. R. Bertolino, and M. D. Lang. 1979. The Japanese longline fishery in the Gulf of Mexico, Mar Fish. Rev. 41:2.3-28. Lutcavage, M. E. 1996. Planning your next meal: leatherback travel routes and ocean fronts. In J. Keinath. D. Barnard, .J. A. Musick, and B. A. Bell (compilers), Proceedings of the fifteenth annual symposium on sea turtle biology and conservation, p. 174-178. U.S. Dep. Commer, NOAA Tech. Memo. NMFS-SEFSC-.387. Lutcavage, M., and J. A. Musick. 1985. Aspects of the biology of sea turtles in Virginia. Copeia 1985:449-4.56. Maul, G. A., F. Williams, M. Roffer, and F. M. Sousa. 1984. Remotely sensed oceanographic patterns and vari- ability of bluefin tuna catch in the Gulf of Mexico. Oceanol. Acta 7:469-479. Morreale, S.J., E. A. Standora, F. V. Paladino, and J. R. Spotila. 1994. Leatherback migrations along deepwater bathy met- ric contours. In B. A. Schroeder and B. E. Witherington (compilers). Proceedings of the thirteenth annual sympo- sium on sea turtle biology and conservation, 109- 110. U.S. Dep. Commer., NOAA Tech. Memo. NMFS- SEFSC-341. [NTIS Accession No. PB95239554/11,| Podesta, G. P., J. A. Browder, and J. J. Hoey. 1993. Exploring the association between swordfish catch rates and thermal fronts on L'.S. longline grounds in the western North Atlantic Ocean. Continental Shelf Res. 13:253-277. NOTE Witzell: Distribution and abundance of sea turtles caught in the western Atlantic Ocean 211 Pritchard, P. C. H. 1971. The leatherback or leathery turtle, Dermochelys corlacea. Int. Union Conserv. Nat.. lUCN Monogr. 1, 39 p. Sakagawa, G. T., A. L. Coan, and N. W. Bartoo. 1987. Patterns in longline fishery data and catches of big- eye tuna, Thunnus uhesus. Mar. Fish. Rev. 49:.57-66. Schmitz, W. J., and M. S. McCartney. 1993. On the North .Atlantic circulation. Reviews in Geo- physics 31:29-49. Schroeder, B. A., and N. B. Thompson. 1987. Distribution of the loggerhead turtle, Caretta caretta, and the leatherback turtle, Dermochelys coriacea, in the Cape Canaveral, Florida area: results of aerial surveys. /;; W. N. Witzell led. i. Ecology of East Florida sea turtles: pro- ceedings of the Cape Canaveral, Florida, sea turtle work- shop, p. 45-.53. U.S. Dep. Commer., NOAA Tech. Rep. NMFS 53. Shoop, C. R., and R. D. Kenney. 1992. Seasonal distributions and abundances of loggerhead and leatherback sea turtles in waters of the northeastern United States. Herpetol. Monogr 6:43-67. Wiebe, P. H., and T. J. McDougall (eds.). 1986. Warm core-rings: studies of their physics, chemistry and biology. Deep Sea Res. 33:1455-1922. Witzell, W. N. 1984. The incidental capture of sea turtles in the Atlantic U.S. Fishery Conservation Zone by the Japanese tuna longline fleet, 1978-1981. Mar Fish. Rev. 46:56-58. 212 Fishery Bulletin 97(1), 1999 Superintendent of Documents Publications Order Form *5178 I I YES, please send me the following publications: Subscriptions to Fishery Bulletin for $35.00 per year ($43.75 foreign) The total cost of my order is $ . . Prices include regular domestic postage and handling and are subject to change. (Company or Personal Name) (Please type or print) (Additional address/attention line) (Street address) (City State, ZIP Code) (Daytime phone including area code) (Purchase Order No.) Charge your order. ITS EASY! VISA Please Choose Method of Payment: I I Check Payable to the Superintendent of Documents □ GPO Deposit Account | | | | | | | l-H I I VISA or MasterCard Account (Credit card expiration date) (Authorizing Signature) Mail To: Superintendent of Documents RO. Box 371954, Pittsburgh, PA 15250-7954 To fax your orders (202) 512-2250 Thank you for your order! U.S. Department of Commerce Seattle, Washington Volume 97 Number 2 April 1999 Hshery Bulletin Contents ^'^^6 1999 The National Marine Fisheries Service iNMFSi does not approve, recommend. or endorse any proprietary product or proprietary material mentioned in this pubhcation. No reference shall be made to NMFS, or to this publication furnished by NMFS, in any advertising or sales promotion which would indicate or imply that NMFS approves, recommends, or endorses any proprietary product or proprietary material mentioned herein, or which has as its purpose an intent to cause directly or indirectly the adver- tised product to be used or purchased because of this NMFS publication. 213-226 227-242 243-255 256-263 264-272 273-281 282-293 294-305 306-319 A/tides Cooper, Andrew B., and Marc Mangel The dangers of Ignoring metapopulatlon structure for the conservation of salmonlds Francis, Malcolm P., Kevin P. Mulligan, Nick M. Davies, and Michael P. Beentjes Age and growth estimates for New Zealand hapuku, Polyprion oxygeneios Goodyear, C. Phillip An analysis of the possible utility of time-area closures to minimize blllfish bycatch by U.S. pelagic longllnes Heifetz, Jonathan, Delsa Anderl, Nancy E. Maloney, and Thomas L. Rutecki Age validation and analysis of ageing error from marked and recaptured sablefish, Anoplopoma fimbria Krieger, Kenneth J., and Daniel H. ito Distribution and abundance of shortraker rockfish, Sebastes borealis, and rougheye rockfish, 5. aleutianus, determined from a manned submersible MacFarlane, R. Bruce, and Elizabeth C. Norton Nutritional dynamics during embryonic development in the viviparous genus Sebastes and their application to the assessment of reproductive success Macy, William K. Ill, Ann G. Durbin, and Edward G. Durbin Metabolic rate in relation to temperature and swimming speed, and the cost of filter feeding in Atlantic menhaden, Brevoortia tyrannus Miller, Thomas J., Tomasz Herra, and William C. Leggett The relation between otolith size and larval size at hatching for Atlantic cod, Gadus morfiua Quiiionez-Velazquez, Casimiro Age validation and growth of larval and juvenile haddock, Melanogrammus aeglefinus, and pollock, Pollactiius virens, on the Scotian Shelf Fishery Bulletin 97(2), 1999 320-331 Stanley, Richard D., Robert Kieser, Bruce M. Leaman, and Ken D. Cooke Diel vertical migration by yellowtail rockfish, Sebastes flavidus, and its impact on acoustic biomass estimation 332-339 Suryan, Robert M., and James T. Harvey Variability in reactions of Pacific harbor seals, Phoca vitulina richardsi, to disturbance 340-361 Terwilliger, Mark R., and Thomas A. Munroe Age, growth, longevity, and mortality of blackcheek tonguefish, Symphurus plagiusa (Cynoglossidae: Pleuronectiformes), in Chesapeake Bay, Virginia 362-371 Thompson, Bruce A., Marty Beasley, and Charles A. Wilson Age distribution and growth of greater amberjack, Seriola dumerili, from the north-central Gulf of Mexico Notes 372-379 Adams, Douglas H., and Robert H. McMichael Jr. Mercury levels in four species of sharks from the Atlantic coast of Florida 380-386 Beal, Brian F., Robert Bayer, M. Gayle Kraus, and Samuel R. Chapman A unique shell marker in juvenile, hatchery-reared individuals of the softshell clam, Mya arenaria L. 387-391 Carlson, John K. Occurrence of neonate and juvenile sandbar sharks, Carcharhlnus plumbeus, in the northeastern Gulf of Mexico 392-395 Holland, Kim N., Pierre Kleiber, and Stephen M. Kajiura Different residence times of yellowfin tuna, Thunnus albacares, and bigeye tuna, T. obesus, found in mixed aggregations over a seamount 396-401 O'Neill, Michael F., David J. Die, Brian R. Taylor, and Malcolm J. Faddy Accuracy of at-sea commercial size grading of tiger prawns (Penaeus esculentus and P. semisulcatus) in the Australian northern prawn fishery 402-405 Vidal, Erica A. G., and Manuel Maimovici Digestive tract parasites in rhynchoteuthion squid paralarvae, particularly in ///ex argentinus (Cephalopoda: Ommastrephidae) 406-409 Yang, Mei-Sun, and Benjamin N. Page Diet of Pacific sleeper shark, Somniosus pacificus, in the Gulf of Alaska 410 Subscription form Papers for Fishery Bulletin 97(2) were reviewed and accepted for publication under the direction of the previous scientific editor. Dr. John Pearce, and his editorial assistant, Laura Garner. 213 Abstract.— Because of their tendency to return to natal streams, salmonid populations have often been viewed in ecological isolation, although the notion of an evolutionarily significant unit (ESU) recognizes dispersal on evolu- tionary time scales. We investigated the consequences of dispersal ( straying) on an ecological time scale where straying creates a metapopulation structure for salmonid streams within an ESU. We developed a simple model for salmonid metapopulations. focusing on source and sink populations, and used the model to highlight the dangers of ignor- ing this structure in conservation ef- forts. We show that exactly the wrong conservation efforts may occur if metapopulation structure exists but is ignored. The dangers of ignoring metapopulation structure for the conservation of salmonids Andrew B. Cooper Quantitative Ecology and Resource Management University of Washington Box 357980 Seattle, Washington 98195-7980 E-mail address andy acqs Washington edu Marc Mangel Department of Environmental Studies and Institute of Marine Sciences University of California Santa Cruz, California 95064 Manuscript accepted 5 Mav 1998, Fish, Bull. 97:213-226 1 1999), The U.S. National Marine Fisher- ies Service (NMFS) Status Review of Coho Salmon (Oncorhynchus kisutch ) from Washington, Oregon, and California (Weitkamp et al., 1995) formalized the agency's ap- proach to defining both the bound- aries and conservation status of dis- tinct segments of salmonid popula- tions for potential listing under the Endangered Species Act. With this approach, originally developed by Waples (1991), a population or group of populations is considered distinct if they are". . . substantially reproductively isolated from conspe- cific populations," and if they are considered ". . . an important com- ponent of the evolutionary legacy of the species" (Weitkamp et al., 1995, p. 3). A distinct population or group populations is referred to as an evo- lutionarily significant unit (ESU) of the species. For a group of populations to be classified as an ESU, the popula- tions must be reproductively iso- lated from other populations but not from each other. An ESU also im- plies successful dispersal and repro- duction between populations on an evolutionary time scale. A meta- population is a group of populations (demes) linked by dispersal of indi- viduals on a shorter ecological time scale such that dispersal affects both the genetics of the individual demes and their abundance and dynamics (Levins, 1969; Ruxton, 1996; Ruxton and Doebeh, 1996). To maintain consistency with the eco- logical literature, we have used the term "metapopulation" to refer to the group of populations or demes, and the term "population" or "deme" (often used interchangeably, see Policansky and Magnuson, 1998) to refer to one of the individual popu- lations that make up the metapopu- lation. ESUs and metapopulations overlap on the continuum of popu- lation structures. Although a meta- population will always compose part of, or the entire, ESU, an ESU does not have to contain any meta- population structure. Dispersal be- tween demes within a metapopu- lation must be great enough to af- fect the dynamics of the demes and the recolonization of habitats of ex- tinct demes. Within an ESU, dis- persal must only be great enough to contribute to the genetic make-up of component populations; it does not have to lead to recolonization events or affect the population dynamics. Salmon typically return to their natal streams to spawn. However, 214 Fishery Bulletin 97(2), 1999 some individuals stray to streams other than their natal one, and those streams may be inhabited or uninhabited by other conspecifc populations (Ricker, 1972; Quinn, 1993). Indeed, straying constitutes the process by which salmonids colonize new habitats (Milner and Bailey, 1989; Wood, 1995). Individuals that stray during the spawning migration may thus serve as the mechanism for dispersal between salmo- nid populations on both an evolutionary and ecologi- cal time scale. Reliable estimates of the magnitude of straying are rare and span a wide range of values across and within species (Quinn, 1993; Pascual and Quinn, 1994). Quinn and Fresh ( 1984) documented a stray- ing rate of 1.4% in their study of wild chinook salmon (Oncorhynchus tshawytscha ) from the Cowlitz River Hatchery, Washington. Quinn et. al(1991) estimated hatchery straying rates ranging from 9.9 to 27.5% for five populations of autumn chinook on the Co- lumbia River. Heard (1991) estimated that, in gen- eral, nearly 10% of wild pink salmon (Oncorhynchus gorbitscha) stray from their natal streams. Labelle (1992) estimated that approximately 4.7% of indi- vidual coho salmon strayed between nine separate streams along the coast of Vancouver Island, British Colombia, but that straying could be greater than 40% for some streams in some years. Genetic stud- ies such as that of Gall et al. ( 1992) suggest that the average number of migrants exchanging genes per generation (Nm) in west coast Chinook salmon popu- lations is on the order of 5-15 individuals. Because the incidence of straying is common and the magnitude of straying is so variable, it is quite likely that metapopulation structure could exist for at least some salmonid populations. In fact, the Na- tional Research Council's report on Pacific Northwest salmonids recognizes that ". . . maintaining a metapopulation structure with good geographic dis- tribution should be a top management priority to sustain salmon populations over the long term" (NRC, 1996, p. 8). Given the geographic scale of the straying documented in the previous studies, com- pared with the range of an ESU (note that the entire West Coast comprised only six ESUs), it is possible that a single ESU may even contain multiple metapopulations, as would be expected because ESUs are explicitly evolutionary constructs, whereas metapopulations are explicitly ecological constructs. In this paper, we investigate the implications of metapopulation structure for conservation efforts given a variety of spatial scal^s. In particular, we identify the problems such structure could cause for managers if it left undetected. If one is concerned strictly with the risk of extinction for a species, metapopulation structure may be quite beneficial (Levins, 1970; Hanski and Gilpin, 1991; Hanski, 1994; Ruxton, 1996; Ruxton and Doebeli, 1996). Be- cause the metapopulation occurs in patches (each of which contains a deme with its own probability of extinction) and because these demes are connected through dispersal, if any single deme becomes ex- tinct, then there is a nonzero probability that the patch will be recolonized by individuals from another deme. Over time, an individual patch may therefore experience multiple extinctions and recolonization events. These events result in the metapopulation as a whole persisting far longer than any one of its individual demes. Potential problems arise when one is concerned not just with the risk of extinction but with the management (and therefore monitoring) of these populations. In the most simple metapopulation model, one as- sumes that all demes, and the patches they inhabit, are identical (Levins, 1970). This, however, need not be the case, and in the real world, is likely not to be the case. One metapopulation model that takes such variation into account is the source-sink meta- population model (Pulliam, 1988). In this model, sink habitats are patches where local mortality exceeds local reproduction (so that R^^1 or r>0). Populations in source habitats can persist without the populations in the sink habitats, but the opposite is not true. There are no assumptions regarding the relative abundance of individuals between these source and sink patches. In fact, it is quite possible for the sink patches to have larger populations than source popu- lations (Pulliam, 1988). For example, if competitively dominant individuals hold territories of fixed size (as is the case with some bird species), a source habitat would be highly productive, yet would have a con- stant population size because all subdominant indi- viduals would be forced to disperse. If this dispersal rate into the sink habitats were greater than the natural rate of decline (the difference between births and deaths) in the sinks, then sink habitats could contain more individuals than the source habitats. In such a case, undetected metapopulation structure could lead managers astray. When metapopulation structure (especially source- sink dynamics) exists, the abundance of a species in an area can be disconnected from the specific survi- vorship and fecundity rates of that area owing to the effects of immigration. If ignored, this disconnection poses two problems for managers, both of which are made worse if the jurisdiction of the manager does Cooper and Mangel: Metapopulation structure in the conservation of salmonlds 215 not cover the complete metapopulation. First, if man- agers are looking strictly at the abundance of indi- viduals, they could be lulled into a false sense of se- curity. The size of the demes in the source and sink habitats could be relatively constant despite the fact that, without the demes in the source, the demes in the sink would become extinct. Brawn and Robinson (1996) uncovered this very scenario with Neotropi- cal migrant birds in Illinois. The second problem is even more insidious. If deme abundance is no longer a good indicator of habitat quality, managers could be led into conserving the wrong type of habitat (van Home, 1983; Pulliam, 1988). Gowan and Fausch ( 1996) demonstrated how this could occur regarding the effects of habitat changes on the demography of a variety of trout spe- cies in Colorado, although they did not discuss their results in terms of metapopulations. Over an eight- year period (four generations of trout), Gowan and Fausch ( 1996) discovered that the addition of woody debris in treatment areas significantly increased the number of individuals and the total trout biomass in treatment areas in relation to the control areas. How- ever, with the aid of fin marks (clipped fins) and in- dividual tags, they discovered that survival, indi- vidual growth, and recruitment rates in the treat- ment areas were not significantly different from those in the control areas. Immigration from outside the study area to the treatment sites was solely respon- sible for the increase in abundance and total biom- ass. If Gowan and Fausch (1996) had not been able to account for immigration to the site, they would likely not have been able to discern the true effects of the addition of woody debris and would have likely mistaken increased density for increased habitat productivity (cf Hunter, 1991 ). Although source-sink metapopulation structure was not the cause of these results, such an example demonstrates how reliance on abundance or density estimates can lead manag- ers astray when immigration or emigration is not taken into consideration. Those faced with the responsibility of managing salmonid populations may encounter these very prob- lems and issues. In the remainder of this paper, we develop a model to help focus ideas about the poten- tial dangers of undetected metapopulation structure for salmonid conservation. Materials and methods The model The model is simple, and the form of the model was chosen for ease of comprehension. We found that even such a simple model was adequate to illustrate the possible consequences of ignoring metapopulation structure. We considered a group of generic salmonid demes that reside in streams that are distributed evenly along some waterway but that are close enough so that straying between any of the two groups is pos- sible (though not necessarily with equal probability). The scale was completely generic. The streams could be tributaries to a single river, rivers within a wa- tershed, or even separate watersheds. Next, we num- bered these streams consecutively along this water- way. Each deme was then indexed by the number associated with the stream in which it resides (e.g. deme 4 resides between deme 3 and deme 5 along this waterway). For computational purposes, we con- sidered 10 streams, which is equivalent to a metapopulation consisting of 10 demes spread over 10 patches. Assuming that density-dependent effects could be ignored (which, except for AUee effects, would be the case for any recovering population), the fundamental variables are N{i,t) = the deme abundance in stream ( in year t; r{i,t) = the per-capita reproduction in stream / in year t; s(J,i,t) = thenumber offish that stray from their na- tal streamy to stream / in year t; and f = the fraction of fish that stray from their na- tal stream (assumed equal for all demes). For simplicity's sake, we assumed that strays have the same reproductive potential as nonstrays in a given stream. This assumption decreases the param- eter space but does not affect overall dynamics of the model. It does, however, limit the direct applicabil- ity of our specific examples to streams that are rela- tively close in proximity, yet, as will be explained, does not diminish the danger for management at the watershed, basin, or even ESU level. By incorporat- ing an additional parameter to account for the dif- ferential reproductive potential, some of the dynam- ics would simply have been dampened, making them more difficult to perceive. Therefore, the population dynamics for a deme are NiiJ + 1} = r(i,t)\N{i,t)a- f) + ^s(j,i,tn . (1) As with differential reproduction, the assumption that there are more complicated population dynam- ics (e.g. Ricker stock-recruitment relationships) would not change the basic message of our paper but would make it harder to perceive. In a more compli- 216 Fishery Bulletin 97(2), 1999 cated model, or one designed for a nongeneric salmo- nid, the model would be indexed by generation time instead of year, but the potential problems described in this paper would still apply. We assumed s(J,i,t) is an exponential function of the distance between stream ; and j (see Hanski, 1994). We assumed that the streams are evenly spaced along the waterway and numbered consecu- tively, so that the distance between stream / and streamy is proportional to | i-j\ . Therefore, the num- ber of individuals straying from stream j to stream / is s(j,i,t)- N(j,t)fe -m\i-j\ 1' -m\k-j\ (2) where m defines the rate at which straying decreases with distance, and the denominator is a normaliza- tion so that all strays end up in a stream (i.e. none are lost to the system). We assumed that per-capita reproduction rate of each deme is a function of some baseline rate of per- capita reproduction that is equal for all streams (e.g. ocean conditions and harvest) plus a function of a component of the habitat that contributes positively towards per-capita reproduction (e.g. width of the riparian zone) and a function of a component of the habitat that contributes negatively to per-capita re- production (e.g. road density). If Zq is the baseline per-capita rate of reproduction and z^{i,t) and z^^i.t) are the amounts of the beneficial and detrimental habitat components for stream / in year t, then the per-capita rate of reproduction for stream / in year t is modeled as /•(,.n = zo + g{i-e--"'"}-6{i-e--"-"} (3) where g = the maximum increase in the per-capita reproduction due to the beneficial habi- tat component; and b = the maximum decrease in the per-capita reproduction due to the detrimental habitat component. As z^(i,t) mcreases, e -zhi.t) 0, so that the effect of the beneficial habitat component approaches an as- ymptote at g. Similarly, the detrimental habitat component approaches an asymptote at b. There- fore, r{i,t] is constrained to lie between z^-b and Zf^+g. When r{i,t)il-P >1, the deme is a source popu- lation (rate of change due to per-capita reproduction counteracts the rate of change due to emigration); when r(i,t) ( 1-/1 <1, the deme is a sink population ( it can not sustain itself without immigration from other streams). To include the effects of temporally varying envi- ronments, we added '27tt' qsm to the per-capita reproduction. This causes per-capita reproduction to change sinusoidally with a maximum change o{2q with a ti'-year period. Such oscillations could be due to events such as El Nino (Pearcy, 1992) and essentially represent changes in the baseline conditions (z^) over time. Simulations We set the baseline per-capita rate of reproduction with ^Q=l,g=0.2, and 6=0.15. Thus per-capita repro- duction was constrained to 0.85 < r(i,t) <1.2. We drew the Zj(/,0) and z.,(z.O) from a gamma distribution with parameters 1 and 1 (Hilborn and Mangel, 1997). We set f- 0.05 for all populations. This value lies within the ranges found by most of the previously mentioned research on salmonid straying rates. Streams were labeled as sources and sinks on the basis of their initial per-capita rate of reproduction given this straying rate. Therefore, a source popula- tion was one with an initial per-capita rate of repro- duction greater than or equal to 1.05, and a sink population was one whose rate was less than 1.05. With these parameters, about 40% of the streams were sources and 60*^ were sinks, as would be the case for a heavily impacted region (Fig. 1 ). We set m =0.1. The initial deme abundance for each stream was assumed to be proportional to the initial per-capita reproduction rate for that stream, even though such relationships may not hold over time (van Home, 1983). As such, the initial deme abundance, Nii.O), was set equal to 100r(;,0). We simulated each metapopulation over a 100-year period, using four sce- narios with at least 150 replications for each scenario: 1 All parameters were constant over the 100-year period; 2 Starting in year 5, for all initial source popula- tions {r{i,0)> 1.05), the good habitat component (2j) decreased by 5'^ of its value from the year before, and the bad habitat component (z^) in- creased by 57r of its value from the year before. 3 Same as scenario 2, except that all habitats with /•((,0)>1 were degraded. 4 Temporally varying environment was incorpo- rated into scenario 2. In this case, the baseline per-capita rate of reproduction (z,,) oscillated be- tween 0.95 and 1.05 over a 20-year period. Cooper and Mangel: Metapopulation structure in the conservation of salmonids 217 0 06 005 ^ ^ 004 / frequency s / <1> > to ■5 0 02 y SINK SOURCE \ DC y (62.5%) (37.5%) ^V 001 / \ r) V m ic h- Net growth rates Figure 1 Distribution of net growth rates. Our model generates variation in per-capita growth rates in local populations. A population is a source if its per-capita growth rate exceeds 1.05 and is a sink otherwise. Results When habitats are constant over the simulation pe- riod, exponential growth occurs DS'vf of the time. Because no density dependence was incorporated in the model, denies in the source habitats increased exponentially, as did the number of strays from these sources. Such a situation could occur if, for instance, harvesting pressure was decreased on a deme that had been severely overharvested in the past (lead- ing to low abundance) but whose habitat was rela- tively pristine (maintaining high productivity). This results in all dynamics in the sink populations being obscured by the massive number of immigrants. It is possible that all the streams constitute sinks (all H;,0) < 1.05); this occurred in 1.3% of the simu- lations. In this case, there is still the possibility that the metapopulation as a whole could persist for de- cades before all populations began to decrease (Fig. 2). The reason for this persistence is that individu- als that stray are lost only to their natal stream, not to the metapopulation as a whole. The case shown in Figure 2 results when the losses due to some habi- tats (where r(/,nl). Forty percent of the all-sink metapopulations produced populations that did not decrease over the course of the 100-year simulation. If we had incorporated a parameter to represent decreased reproductive success of strays, the incidence of nondecreasing, all-sink meta- populations would have been lower. In scenario 2 (Fig. 3), the source populations (demes 3, 4, and 5) initially increased exponentially but eventually began to decrease as the habitat deg- radation increased. Habitat degradation leads to peaks in local deme abundance (Fig. 4A). The mode of this distribution was around year 20 despite the fact that habitat degradation began in year 5. Fur- thermore, in over 3% of the sources, deme sizes in- creased throughout the 100-year simulation. The population trajectories for the sink habitats were less intuitive ( Fig. 3 ). None of their habitat com- ponents changed, yet some demes increased (deme 1) or stayed constant (deme 9) over a number of decades, whereas others decreased. The result depends on the proximity to sources (noting that deme number trans- lates to the location of the deme along the water- way), the per-capita rate of reproduction in those sources, and the sink's own per-capita rate of repro- duction. Over 'd'^c of the sink habitats never attained Fishery Bulletin 97(2), 1999 Deme 1,r(1,0) = 1.00 Deme2, r(2.0) = 1 03 150 - loo- 150 - 100 - se ' ^^^ ' 50- n - 1 *- o oooooooo O o o o o o o o o Deme3, r{3,0) = 0.92 OJ CO TT in -. . 50 0 ' — _ ^ ^ o o o o o o o o o cjcOTfincor^oooj Deme 5, r(5.0} = 0,98 CM n -^ in CD r^ GO Deme6. r(6,0) = 1.02 Ol ® N W c 150 T 1M) n o ' ro 100 ^ ^ 100 % 50^ 0. " — -^ 50 0 ■ — 1 ^ o oooooooo ^ o o o o o o o o o ojco-^intor^coo) Deme 7. r(7,0) =0.92 c\j en TT in CD r- CO DemeS. r(8.0) = 0.87 O) 150 T 150 -| 100 - s. 100 50 0 ^ 50 0 CMCOTTLntDNOOO Deme 9. r{9.0) = 0,97 c\j n -^ in CD h- CO Deme 10. r(10,0) = 0 99 150-, 150 100 .^ 100 50 50 0 ' ■ o CMfD-^LncDh-COfJj Year (M CO -rj- in CD h- CO O) Figure 2 The dyna mics of the metapopulation when all local populatio ns are sinks. With scenario 1 or 2 (see "Simulat ons" section), metapopulation structure and disperse il actually lead to an increase in some populatio n sizes over the short term, before they decline deme sizes larger than their initial deme size, but nearly 7% of the demes in sink habitats continued to increase over the entire simulation (Fig. 4B). The fact that deme sizes in some sink habitats do not decrease over the simulation period is a result of the time of the simulation in relation to the rate of habitat deg- radation in the sources. If the simulations had been longer, all habitats would eventually have reached Cooper and Mangel: Metapopulatlon structure in the conservation of salmonids 219 Deme1,r(1,0)=1.03(sink) Deme 2, r(2.0)=0.94 (sink) 200 - 200 - 100 * ' 100 ^ . ^-ooooooooo -oooooooo o Deme 3. r(3.0)=1 , 1 0 (source) Deme 4, r(4,0)=1.10 (source) 200 /~\ 200- /~\ 100 ^ \.^_^ 100- ■ \^_ ^ooooooooo T-c\jco^in;or^co g Deme 5, r{5.0)=1 07 (source) Deme 6, r(6,0)=0.97 (sink) o N m 200 1 100 ^ 0- - 200- - ^^\^_ 100 - ~.-_____^ 1 1 1 1 1 1 1 O i-OOOOOOOOO T- Q- •-cvjcO'TintOf^oocj) oooooooo o Deme 7. r(7,0)=0.98 (sink) Demee, r{8.0)=0.91 (sink) 200 [ 200 100 100 <^_______ n ^ ^ooooooooo T- oooooooo 8 Deme9. r(9.0)=1.00(sink) Deme 10. r(10.0)=0.92 (sink) 200 200 100 100 _ '-ojcoTrintor^coO) oooooooo 8 Year Figure 3 The dynamics of the local populations when some are sources a nd some are sinks. With scenario 2 (see "Simulations" section), as habitat degradation proceeds (startir ig in year 5), habitats for source popula- | tions slowly become degraded and end up as sinks. the state of being sink habitats, which would then have resulted in only a few instances of sustained or increasing metapopulations (as described above). Because the number of migrants from one deme to another decreases with distance, the year in which the maximum deme size is attained by the sink de- creases as the distance from a source increases; thus the sinks rely on source populations for their viabil- 220 Fishery Bulletin 97(2), 1999 cr 0.30 0.20 0.10 0,00 J IIIIIIMIIIHiMIMIIIIMfWWTfWW ^ CD ^ cy CO CO Y- CD »- (D in in CD CD 00 00 o 0,30 0,20 0,10 0.00 1- CD T- (D 1- CD '- CD 1- CD ^ CD ^ in ID CD CD 03 02 01 0 C\J CM CO CO mill r iniiiiMii T- CD T- CD in lo CO to Year Figure 4 Proportion of sources (A and C) and sinks (B and D) attaining their maxi- mum population size in a given year for scenario 2 (A and B) and scenario 3 (C and D). (A) Habitat degradation causes source population size to peak some- time after year 5, although a few populations maintain themselves through- out the 100-year simulation period. (Bl About lO'/c of the sink populations at- tain maximum size at the start of the simulation. A few attain maximum size at the end of the simulations, but most attain maximum size around year 40. They are supported by source populations. (C) The frequency distribution of peaks in population size for populations with initial per-capita growth rate >1. Compare with Figure 4A. (Dl Most populations with initial per-capita growth rate <1 attain peak size at the .start of the simulated period. Compare with Figure 4B. ity (Fig. 5). The variance about these points is due not only to the per-capita reproductive rate of the sink and its nearest source but also to a failure to account for the location of any qther sources that may be of equal or greater distance from the nearest source and thus contribute to the sink's dynamics. When all demes with initial per-capita growth rates greater than one are affected by habitat degradation. declines in deme size occur more frequently, more rapidly, and sooner than in the previous case (Figs. 4 and 6) In this scenario, 100'^ of the demes were driven towards extinction. When an oscillating environment is incorporated, the general trends in abundance are similar to those in scenario 2, but now each deme also tracks the en- vironment (Fig. 7). All sources (demes 3, 5, and 7) Cooper and Mangel: Metapopulation structure in the conservation of salmonids 221 60 50 - ■ 40 > 30 I < ' ^ , 20 " 10 J J J 0 Distance Figure 5 Average year in which sinks attained their maximum population size as a function of the distance to the nearest source. The year in which a sink population attains its maximum size depends upon the distance to the closest source, but there is considerable variation in dependence. Error bars represent one standard deviation. initially increase but then decrease as habitat deg- radation progresses. Some demes in sink habitats, such as demes 1 and 4, were relatively stable if not increasing over a fair portion of the simulation pe- riod before declining towards extinction. Discussion Undetected metapopulation structure in salmonid populations may obscure the signals that managers use to determine the need for convervation action. Abundance trends, either absolute numbers or simple indices of abundance, constitute the primary input into analyses used for fisheries management and decision-making (Hilborn and Walters, 1992). The majority of these techniques assume that the popu- lation in question is a closed system, that any immi- gration or emigration can be considered negligible. In California, Oregon, and Washington, estimates of population or run size for most salmonid species are commonly based on number of spawning fish or redd counts taken from index reaches or streams ( WDFW, 1994; ODFW, 1995; Weitkamp et al., 1995). The absolute numbers of spawning fish or redds counted in these index streams, which incorporate only a minute portion of the available spawning habi- tat within the watershed, are then used to extrapo- late watershed- or basin-level abundance estimates, the very estimates upon which managers base their decisions. In fact, in their discussion on the data available for assessing the population size and risk of extinction for coho salmon along the west coast of the United States, Weitkamp et al. (1995, p. 106) stated ". . . where [stream] surveys were conducted, they are the best local indication we have of popula- tion abundance trends." Index reaches and streams such as these are precisely the situation modeled in the specific examples discussed in this paper; stray- ing between these reaches is likely and differential reproductive success between strays and natal fish may not occur. The NMFS's analysis of coho salmon populations considered short- and long-term trends in abundance to be the main indicators for the risk of extinction but avoided using estimates based on index streams (Weitkamp et al., 1995). Using the trends in abundance for an entire ESU, Weitkamp et al. (1995) attempted to avoid the problems associated with the 222 Fishery Bulletin 97(2), 1999 Deme 1, r(1.0) = 0.95 (sink) Deme2, r(2.0)=1 11 (source) 300 ^^— .^ 200 ■ 200 / N. 100 ^ 100 \.^ '-ooooooo '- oj CO ^ IT) CO r* OO ■r-OOOOOOO a> Oi ^ojco'^tncoh- o CO o Oi Deme 3, r(3,0) = 1.10 (source) Deme4, r(4,0) = 0.94(sink) '^nA 300 200 - -/^ "\ 200- ■ 100- \^^ 100 ■ ^ooooooo »~ CM CO ^ in CO f^ OO ■.-OOOOOCDO COO) ^cxjcOTfiDiOh- o CO o o Deme 5. r(5.0) = 0.96 (sink) Deme 6, r(6.0) = 1 .06 (source) o c o 200 - 200 - ^^__^^ Q. O CL 100 J 100 ^"^.'..^^^ ■ — ' ooooooo ■^ csj CO -^ tn o h- O O CO Oi '- Cy CO TT ID to r^ § Deme 7. r(7,0) = 1.03 (sink) Deme 8, r{8,0) = 1.07 (source) 200 ■ 200 . ^v^ 100 ^ "^^^--..^^^ 100 ^ ^^..^^ »-ooooooo T- CM CO -^ m CD r^ C0C3) ^CMCO'^tmcDh- g s Deme 9, r(9.0) = 0.93 (Sink) Deme 10, r(10,0) = 1.03 (sink) 200 200 100 ^ 100 ' ^--^^_^ — ooooooo T- cy CO ^ in CO f^ OO ••-OOOOOtSO COO) v-CMco^inor^ Year Figure 6 o CO o The local population dynamics when all population.^ with initial per-capita growth rate >1 are subject | to h abitat degradation (scenario 3 in ' Simulations" section). Compare with Figure 3. assumption of a closed-system. However, depending on the time scale of the data available and the decrease in reproductive success for strays, undetected metapopu- lation structure could still cause problems. The second problem arises because metapopulation structure can disconnect the abundance-habitat quality relationship. This could lead managers to make erroneous inferences regarding the habitat requirements for a species. The Cooper and Mangel: Metapopulation structure in the conservation of salmonids 223 300 300 200 100 o D. 300 200 100 Deme1,r(1,0)=1.01 (sink) Deme 3, r(3,0) =1.06 (source) Deme 5, r(5,0) =1.06 (source) Deme 7, r(7.0) =1.09 (source) Deme 9, r(9,0) =0.87 (sink) 300 200 100 300 200 300 200 Deme 2, r(2,0) =0.97 (sink) »- CO en Deme 4, r(4,0) =1 .02 (sink) Deme 6. r(6.0) =0,96 (sink) Deme 8, r(8,0) =0 99 (sink) Deme 10, r(10,0) =0.96 (sink) Year Figure 7 When Zg varies (scenario 4 in "Simulations" section), the local population sizes reflect such variation. Compare with Figure 3. sample metapopulations from the simulations provide demonstrations of how this may occur Figure 3 demonstrates one example where meta- population structure could obscure management sig- nals. Keep in mind, there is neither density depen- dence, observation error, nor stochasticity in the population dynamics, and a deme could represent an index reach (conforming to our uniform reproductive 224 Fishery Bulletin 97(2), 1999 success assumption) or an entire stream (where the uniform success assumption may be less accurate). If a manager had the entire time series data for all of the demes, the problem (though not the solution) would be evident; all demes would have declined significantly. Suppose, however, the manager is currently in year 30, and has only 25 years of data (years 5-30). Over that time period, all demes in sink habitats would be relatively stable or decline only slowly, even though none could sustain themselves without immigrants. In fact, deme 1 had been increasing until year 30. The demes in the source habitats, however, showed the ef- fects of habitat degradation. Seeing these trends, the manager could perhaps stabilize these demes and thus inadvertently save the entire metapopulation. Still, the true risk for the sink populations would be unknown to the manager. The sink populations would appear stable not because they were in good condition but rather because of their interconnectedness with one another and the source populations. Rieman and Mclntyre ( 1995 ) suggested that a simi- lar process is occurring with bull trout {Salvelinus confluentus) populations in Idaho. Although bull trout were found to use small streams, they do so only at a very low frequency. Rieman and Mclntyre (1995) concluded that the presence of trout in these streams may be influenced by habitat preference but that these populations depend on dispersal of indi- viduals from larger streams for their sustainability. The addition of a varying environment clouds the picture even more (Fig. 7). The manager must now distinguish between decreases due to environmen- tal factors and decreases due to factors that can be controlled. Imagine a manager starting in year 1. How can one recognize that the metapopulation is collapsing due to anthropogenic effects and not sim- ply to environmentally induced effects? When would the alarms sound? Probably not until sometime af- ter year 30, when even the most historically produc- tive demes do not begin to increase. Seven out often of the demes could not maintain themselves without immigration, there is neither observation error nor stochasticity, and it would still take over 25 years of habitat degradation before the problem was noticed and the alarms sounded. These problems become even more serious when management boundaries do not coincide with metapopulation boundaries or when only a portion of the metapopulation is used as index streams, as is likely the case. In such a situation, a manager will be concerned with, have jurisdiction over, and maybe have data on some subset of the metapopulation. Returning to Figure 3 (where managers do not have to deal with fluctuating environments), imagine if a manager had responsibility, and therefore data, for only demes 6, 7, 8, 9, and 10, which are nevertheless a significant portion of the metapopulation. Between years 5 and 30, all demes were relatively stable, de- spite the fact that without immigration, even the most productive deme (deme 9) would decrease at a rate of 5% per year. After year 30, all these demes begin to decline. The manager would, of course, be- gin looking at these demes to try and see what was causing this decline; a good manager would try to find what had changed. In fact, nothing has changed with these demes; they have exactly the same rates of per-capita reproduction and straying over the en- tire simulation. Only the number of immigrants has changed. Without looking beyond their own jurisdic- tion or at streams other than the index streams, managers would not find the true cause for the change in dynamics. As the number of demes that are used as index streams or that lie within a manager's jurisdiction decreases, the likelihood of such a problem increases. Considering that over 90'7r of the sink populations increased in size during some portion of the simulation time horizon (i.e. their maximum size was reached sometime after year 1, Fig. 4B), the fate of sink populations has as much, if not more, to do with the health of the demes in other habitats as with the quality of their own habitat. Another problem associated with undetected metapopulation structure arises directly from this last example. A manager may see a change in the dynamics in the populations and look for the cause. However, the cause of this change is outside the manager's jurisdiction or data set; it is not local. If the manager looks only for local causes for this change, she or he is likely to find some variable that is correlated with this cause and possibly infer local causality. Having attributed causality to a local vari- able, the manager would likely start funding projects to fix the perceived problem in the correlated vari- able. If one is lucky, very lucky, this variable might have some relationship to the per-capita reproduc- tive rate and thus improve the situation a bit. This might, then, slow the rate of decline, but because it is not the true cause of the decline, the situation would likely continue to deteriorate. How much of a manager's limited resources might be spent on such activities before the true causal relationship was dis- covered? The study by Gowan and Fausch (1996) suggests that this problem may be occurring with habitat enhancement projects for trout in Colorado; managers promote the addition of woody debris in streams because it increases trout density, even though it does not seem to improve the demographic parameters of the population. Including density dependence (as in Ricker stock- recruitment relationships) and observation or pro- Cooper and Mangel: Metapopulation structure in the consen/ation of salmonids 225 cess uncertainty (Hilborn and Mangel, 1997) in the model will not change any of the main conclusions. These factors will make recognizing the problem even more difficult, as is the case in the real world. To uncover metapopulation dynamics, one must explore two aspects of each population's life history in refer- ence to its habitat: dispersal (in the form of immi- gration and emigration ) and the per-capita reproduc- tive rate (defined by survival and reproduction). Watershed-scale estimates of these rates, however, are not appropriate. Given their reliance on index stream counts, managers must know the rates of immigration, emigration, and reproduction specific to that stream to be able to understand the true dy- namics of that stream. The dispersal of individuals is important to determine the extent of the meta- population structure. The focus should be both on absolute numbers as well as on the effective migra- tion rates that account for differences in survival and productivity between the residents and the migrants. Survival and reproduction estimates (the components of the per-capita reproductive rate) will allow the manager to assess the potential importance of the metapopulation structure (i.e. to define sources and sinks). As long as the risk exists for abundance and density estimates to be disconnected from habitat qual- ity and per-capita reproduction, all the above informa- tion is required to make an accurate assessment of the conservation status of the individual demes and to choose the appropriate management actions. A great deal of time, money, and effort is currently directed toward the conser\'ation and improvement of salmonid populations and their habitats. The NMFS report alludes to the fact that metapopulation structure may exist between some salmonid popula- tions, and the NRC (1996) report lists the mainte- nance of metapopulation structure as one of its most important recommendations. Without investigating the possibility of metapopulation structure, research- ers, managers, and policy makers are setting them- selves up to fall into the traps described above: that of either not seeing a problem that may exist or, if they do see it, not knowing the true causes of such a problem. Acknowledgments The initial phases of this work were supported by a contract from the Southwest Fisheries Science Cen- ter, Tiburon, CA, to MM. We thank Robin Waples and Tom Wainwright for providing a conducive set- ting where we could meet. We thank Michael Healey, Andrew Hendry, and three anonymous reviewers for comments on the manuscript. Literature cited Brawn, J. D., and S. K. Robinson. 1996. Source-sink population dynamics may complicate the interpretation of long-term census data. Ecology 77: 3-12. Gall, G. A. E., D. Hartley, B. Bentley, J. Brodziak, R. Gomulkiewicz, and M. Mangel. 1992. Geographic variation in population genetic structure of Chinook salmon from California and Oregon. Fish. Bull. 90:77-100 Gowan, C, and K. D. Fausch. 1996. Long-term demographic responses of trout populations to habitat manipulation in six Colorado streams. Ecol. Appl. 6:931-946. Hanski, I. 1994. A practical model of metapopulation dynamics. J. Anini. Ecol. 63:151-162. Hanski, I., and M. Gilpin. 1991. Metapopulation dynamics: brief history and concep- tual domain. Biol. J. Linnean Soc. 42:3-16. Heard. W. R. 1991. Life history of pink salmon (Oncorhynchus gorbuscha I. In C. Groot and L. Margolis (eds.). Pacific salmon life his- tories, p. 121-230. Univ. British Columbia Press, Van- couver. B.C Hilborn, R., and M. Mangel. 1997. The ecological detective: confronting models with data. Princeton LIniv. Press, Princeton, NJ, 315 p. Hilborn, R., and C. J. Walters. 1992. Quantitative fisheries stock assessment: choice, dy- namics, and uncertainty. Chapman and Hall, New York, NY, 570 p. Hunter, C. J. 1991. Better trout habitat: a guide to stream restoration and management. Island Press, Washington D.C.. 320 p. Labelle, M. 1992. Straying patterns of coho salmon {Oncorhynchus kisutch I stocks from southeast Vancouver Island, British Columbia. Can. J. Fish. Aquat. Sci. 49:1843-1855. Levins, R. 1969. Some demographic and genetic consequences of en- vironmental heterogeneity for biological control. Bull. Ent. Soc. Am. 15:237-240. 1970. Extinction. In M. Gerstenhaber (ed.l. Some math- ematical questions in biology, p. 77-107. Am. Mathemati- cal Society. Providence. RI. Milner, A. M., and R. G. Bailey. 1989. Salmonid colonization of new streams in Glacier Bay National Park. Alaska. Aquacult. Fish. Manage. 20(2): 179-192. NRC (National Research Council). 1996. Upstream: salmon and society in the Pacific Northwest. National Academy Press, Washington, D.C., 452 p. ODFW (Oregon Department of Fish and Wildlife). 1995. Oregon coho salmon biological status assessment and staff conclusions for listing under the Oregon Endangered Species Act (Commission decision draft). Oregon Dep. Fish Wildlife. Portland. OR, 59 p. Pascual, M. A., and T. P. Quinn. 1994. Geographical patterns of straying of fall chinook salmon, Oncorhynchus tshawytscha (Walbaum), from Co- lumbia River (LTSA) hatcheries. Aquacult. Fish. Manage. 25(suppl. 2):17-30. 226 Fishery Bulletin 97(2), 1999 Pearey, W. G. 1992. Ocean ecology of North Pacific salmonids. Univ. Washington Press, Seattle, WA, 179 p. Policansky, D., and J. J. Magnuson. 1998. Genetics, metapopulations, and ecosystem manage- ment of fisheries. Ecol. Appl. 8(suppl. 1): S119-S123. Pulliam, H. R. 1988. Sources, sinks and population regulation. Am. Nat. 1.32:652-661. Quinn, T. P. 1993. A review of homing and straying of wild and hatch- ery-produced salmon. Fish. Res. 18:29-44. Quinn, T. P., and K. Fresh. 1991. Homing and straying in chinook salmon (Oncorhyn- chus tshawytscha) from Cowlitz River hatchery, Wash- mgton. Can. J. Fish. Aquat. Sci. 41:1078-1082. Quinn, T. P., R. S. Nemeth, and D. O. Mclasaac. 1991. Homing and straying patterns of fall chinook salmon in the lower Columbia River. Trans. Am. Fish. Soc. 120:150-156. Ricker, W. E. 1972. Hereditary and environmental factors affecting cer-' tain salmonid populations. In R. C. Simon and P. A. Larkin (eds.). The stock concept in Pacific salmon, p. 27- 160. H. R. MacMillan Lectures in Fisheries, Univ. Brit- ish Columbia, Vancouver, B.C. Rieman, B. E., and J. D. Mclntyre. 1995. Occurrence of bull trout in naturally fragmented habitat patches of various size. Trans. Am. Fish. Soc. 124: 285-296. Ruxton, G. 1996. Synchronization between individuals and the dynam- ics of linked populations. J, Theor. Biol. 183:47-54 Ruxton, G., and M. Doebeli. 1996. Spatial self-organization and persistence of transients in a metapopulation model. Proceedings of the Royal So- ciety of London (series B) 263:1153-1158 van Home, B. 1983. Density as a misleading indicator of habitat quality. J. Wildl. Manage. 47:893-901. Waples, R. S. 1991. Definition of "Species" under the Endangered Spe- cies Act: application to Pacific salmon. U.S. Dep. Commer., NOAA Tech. Memo., NMFS F/NWC-194, 29 p. WDFW (Washington Department of Fish and Wildlife and the Western Washington Treaty Indian Tribes). 1994. Washington State Salmon and Steelhead Stock In- ventory (S. A. S.S.L). Northwest Indian Fisheries Commis- sion, State of Washington, Dep. of Fisheries, Dep. of Wild- life, Olympia, WA, 212 p. Weitkamp, L. A., T. C. Wainwright, G. J. Bryant, G. B. Milner, D. J. Teel, R. G. Kope, and R. S. Waples. 1995. Status review of coho salmon from Washington, Or- egon, and California. U.S. Dep. Commer, NOAA Tech. Memo., NWFSC-24, 258 p. Wood, C. C. 1995. Life history variation and population structure in sockeye salmon. In J. L. Nielsen (ed. I, Evolution and the aquatic ecosystem: defining unique units in population conservation, p. 195-216. American Fisheries Society Symposium, no. 17: symposium on evolution and the aquatic ecosystem: defining unique units in population conservation, Monterey, CA, 23-25 May. 1994. 227 Abstract.— Po/.vpr/on oxygeneios (fam- ily Polyprionidae) is fished commer- cially and recreationally in the south- ern Indian and Pacific Oceans. Esti- mates of growth rate, age at maturity and recruitment, and longevity are re- quired for fishery management. We used thin otolith sections to age P. oxygeneios from New Zealand, where it is known as hapuku. Growth bands were difficult to count, leading to low counting precision and, for some age groups, to a small between- and within- reader ageing bias. These problems, however, had little effect on the shape of growth curves fitted to length-at-age data. An oxytetracycline injection ex- periment supported our hypothesis of annual deposition of an opaque-hyaline band pair, but further validation is re- quired. Independently derived von Bertalanffy growth curves ( from length- frequency data) and growth-rate esti- mates (from tag-recapture data) for young hapuku agreed well with esti- mates from length-at-age data. Juvenile hapuku are pelagic and most switch to a demersal habitat at around 50 cm total length and at an estimated age of 3^ years. They prob- ably recruit to commercial trawl catches at about the same age. Female hapuku appear to grow slightly faster than males. Both sexes mature at about 10- 13 years. The longevity of hapuku is uncertain, but some individuals prob- ably live longer than 60 years. Age and growth estimates for New Zealand hapuku, Polyprion oxygeneios Malcolm P. Francis Kevin P. Mulligan Nick M. Davies Michael P. Beentjes National Institute of Water and Atmospheric Research P, O Box 14-901, Wellington, New Zealand E-mail address (for M P Francis) m francisaniwacn nz Manuscript accepted 5 May 1998. Fish. Bull. 97:227-242(1999). Polyprion oxygeneios (Schneider, 1801) (family Polyprionidae) is a large demersal fish that inhabits temperate and subtropical waters of the southern Indian and Pacific Oceans (Roberts, 1986; Fig, 1), It supports significant fisheries off the Juan Fernandez Islands, Chile, and off New Zealand and is an impor- tant bycatch in south-east Austra- lian longline fisheries (Johnston, 1983; Pavez and Oyarziin, 1985; Pizzaro and Yaiiez, 1985; Kailola et al,, 1993; Annala and Sullivan, 1997). The closely related wreck- fish, P. at7iericanus (Bloch and Schneider, 1801) is found in the North and South Atlantic Oceans, Mediterranean Sea, southern In- dian Ocean, and south-west Pacific Ocean (Roberts, 1986; Sedberry et al,, 1994) and supports fisheries throughout its range (Sedberry et al., 1994, 1996). The ability to age Polyprion spe- cies is essential for proper manage- ment of their widespread fisheries, because it will enable determination of growth rates, ages at maturity and recruitment, longevity, and natural mortality rates. Previous attempts to age the two species have had limited success. Age and growth off, oxygeneios have been estimated from otolith band counts in New Zealand (McDougall, 1975) and from scale annuli counts in Juan Fernandez Island (Pavez and Oyar- ztin, 1985). A study on the age and growth of P, americanus from the north-west Atlantic is currently underway (Sedberry et al., 1994, 1996; Manooch and PottsM. How- ever, otolith bands have proven dif- ficult to count in both species, scale annuli are unreliable for older fish, and the ageing techniques have not been validated. In New Zealand, P. oxygeneios and P. americanus are managed as a single species-unit in the Quota Man- agement System (QMS) (Paul and Davies, 1988; Annala and Sullivan, 1997), Both species are fished com- mercially and recreationally through- out New Zealand, They are highly sought after, but catches are rela- tively low. Commercial landings of both species combined peaked at 2700 t in 1983-84, but since the in- troduction of the QMS, they have been constrained by quotas to less than 1500 t per year (Annala and Sullivan, 1997), Despite these quo- tas, commercial line and set-net fisheries for Polyprion species have been seasonally important for small inshore fishing vessels in many parts of New Zealand. Annual recreational landings were about 500 t in 1991- 94 (Annala and Sullivan, 1997), This paper addresses age and growth in New Zealand P. oxy- Manooch, C. S., and J. Potts. 1997. Na- tional Marine Fisheries Service, 101 Pivers Island Rd, Beaufort, NC 28516. Personal commun. 228 Fishery Bulletin 97(2), 1999 geneios, hereafter called hapuku. However, our re- sults are also relevant to other populations of P. oxygeneios and, because both species of Polypriou have similar otolith structure, our results should also assist in the interpretation of otoliths of P. ameri- canits. We used thin otolith sections to age hapuku and attempted to validate ages using oxytetracycline (as a time marker) and otolith marginal state analy- sis. We also obtained independent growth rate esti- mates from length-frequency and tag-recapture data for comparison with growth rate estimates derived from length-at-age data. Materials and methods Otolith ageing Sagittal otoliths were obtained from 1400 hapuku collected from waters off New Zealand, from 1979 to 1995. Many otoliths were collected during studies by Johnston ( 1983 ) and Roberts ( 1986 ) and made avail- able to us. Since then, additional otoliths have been collected by us during research trawl surveys and sampling trips aboard commercial line vessels, as well as by Ministry of Fisheries scientific observers aboard commercial line and trawl vessels. Most otoliths were accompanied by information on fish total length (TL, to the centimeter below actual length) and sex. Three subsamples of these otoliths were selected for analysis. The first comprised 44 otoliths collected from five sites between Three Kings Islands and Great Barrier Island (Fig. 2) (hereafter called the "northern" sample). The second consisted of 178 otoliths collected from Cook Strait and Kaikoura ( Fig. 2; "central" sample) and the third con- sisted of 28 otoliths from large hapuku collected from other sites throughout New Zealand ("miscellaneous" sample), selected to boost the number of old fish in the data set. One otolith from each pair was viewed under trans- mitted polarized light, and the straightest dorso- nuclear ridge was marked with a pen to guide the saw for sectioning. Each otolith was then embedded in a block of clear epoxy resin and sectioned trans- versely through the core at a thickness of 650 \xm by using a dual-blade high-speed diamond saw. One face of the section was polished with caz'borundum paper and glued on to a microscope slide with thermoplas- tic cement. The opposite face was then ground and polished to a final thickness of 200-350 |.im, depend- ing on the clarity of the otolith bands when viewed under both reflected and transmitted light. Opaque otolith bands were counted under trans- mitted light in the dorsal part of each section by two readers. Both readers carried out an initial training exercise by making counts on a subsample of the sec- tions, while knowing the collection details of each fish (TL, sex, and site), and then discussed their re- sults. Subsequent counts were carried out "blind"; i.e. the readers did not know the size, sex, or collec- tion site of the hapuku. Reader 1 counted all sec- tions once (Rj) and reader 2 counted all sections twice, the two counts being separated by two months (R., , and R., .,). The readability of sections was scored on a scale from 1 (unreadable) to 5 (exceptionally clear). Otolith band counts were assessed for ageing bias and precision (between readers, and between read- Francis et al.; Age and growth estimates for Polypnon oxygeneios 229 165°E 170° 175° — I 1 1 1 1 1 1 1 1 1 — 180° 35°S 40' 45° Three Kings Is n Otolith site ■ Tagging site Q Otolith and tagging sites Poor Knights Is *a' iGreat Barrier Is Cook Strait Kail were selected as a baseline against which to compare other estimates because it was anticipated that the experience gained during the first reading would produce more reliable age estimates during the second reading. Validation of the annual formation of otolith bands was attempted by marginal increment analysis and oxytetracycline injection. The marginal composition of 76 otoliths from northern and central hapuku, aged 3-8 years, was graded as opaque, narrow hyaline, or wide hyaline. The classification of hyaline zones as narrow or wide was based on a comparison of the marginal zone width with the width of the preceding hyaline zone. A qualitative classification was used because the presence of split opaque bands made it difficult to measure the marginal increment. The data were grouped into two-month intervals because not all months were sampled and because sample sizes were small. Thirty-nine hapuku were tagged (see "Tag-recap- ture" section), injected intramuscularly with oxytet- 230 Fishery Bulletin 97(2), 1999 racycline hydrochloride (OTC), and released at the Poor Knights Islands (Fig. 2). Two OTC brands and dosage rates were used: 27 fish were injected with terramycin at a dosage rate of 80 mg/kg; and 12 were injected with engemycin at 35 mg/kg. Five OTC-in- jected fish were recaptured and their otoliths re- moved. One otolith from each pair was sectioned as described above and viewed under transmitted white light and reflected UV light. The composition of the otolith (number of opaque or hyaline bands) outside the OTC mark was determined, and the amount of new otolith material laid down along the dorso- nuclear axis was measured. Growth curves were fitted to the length-at-age data bu using the von Bertalanffy growth model: L Jl- -KU-tJ ). where L, = the expected length at age t years; L^^= the asymptotic maximum length; K = the von Bertalanffy growth constant; and t,. = the theoretical age at zero length. Length-frequency distributions Hapuku length-frequency distributions were ob- tained from a series of four bottom trawl surveys conducted from the research vessel Tangaroa (tow- ing a 60-mm mesh codend) off Southland (the south- ern coast of South Island and the Stewart Island shelf between latitudes 46-49''S). The surveys were car- ried out annually in February and March, 1993-96, and covered a depth range of 30-600 m. Further de- tails of the surveys are provided in the studies of Hurst and Bagley ( 1994) and Bagley and Hurst ( 1995, 1996a, 1996b). Von Bertalanffy growth curves were fitted to the four length-frequency distributions by using the MULTIFAN model (Fournieretal., 1990). This model analyses multiple length-frequency distributions si- multaneously and uses a maximum likelihood method to estimate the number of age classes repre- sented by the data, the proportions of fish in each age class, and the von Bertalanffy growth param- eters L^ and K. The main assumptions of the MULTIFAN model are the following: 1 ) the lengths of the fish in each age class are normally distributed around their mean length; 2) the mean lengths-at- age lie on or near a von Bertalanffy growth curve; and 3) the standard deviations of the actual lengths about the mean length-at-age are a simple function of the mean length-at-age (Fournier et al., 1990). The growth parameters were estimated by conduct- ing a systematic search across a parameter space of plausible K values (0.005-0.30) and age classes (6- 15). It is possible to constrain the MULTIFAN search further by specifying initial estimates for the mean length-at-age, and the range of the mean length, for one or more age classes (Fournier et al., 1990). How- ever, this is not necessary when the length distribu- tions contain adequate modal structure; therefore such constraints were not used in this study. For each of the identified age classes, MULTIFAN also estimates the ratio of the last to first length standard deviations (S^j and the geometric mean of the first and last standard deviations (S^). The MULTIFAN model was fitted for two different growth hypotheses: 1) constant length standard deviation for all age classes (fitted by setting S^^=l and esti- mating S^); and 2) variable length standard devia- tion across age classes (fitted by estimating both S^ and Sjj). Because all four trawl surveys were con- ducted at the same time of year, the data contain no information on seasonal variability of growth, and no seasonal parameters were fitted. Each combination of K and the number of age classes constituted a model fit. The constant stan- dard deviation hvpothesis was fitted to the data first, followed by the addition of the parameter for variable standard deviation. For each combination of K, num- ber of age classes, and gi'owth hypothesis, the maxi- mum log-likelihood (A) was calculated. Likelihood ra- tio tests were used to test for significant improvement in model fit. Twice the increase in A is distributed as a X~ distribution with degrees of freedom equal to the number of additional parameters. Following Fournier et al. (1990), a significance level of 0.10 was used for testing whether there was any gain in introducing an additional age class in the length-frequency analyses. The test for improvement resulting from the addition of the parameter for variable standard deviation was carried out with a significance level of 0.05. The von Bertalanffy growth parameter t^^ was es- timated from the equation tn t. where a^ = the age estimated by MULTIFAN (in years since zero length) of the youngest age class at the time it first appeared in the length-frequency samples; and t^ - the time elapsed in years between the theoretical birthday and the first ap- pearance of the youngest year class in the samples. The theoretical birthday was defined as 1 Septem- ber based on observations of prespawning fish in June-August and postspawning fish in October- December (Johnston, 1983; Roberts, 1986). Francis et al : Age and growth estimates for Polyprlon oxygeneios 231 Tag recapture 1623 hapuku were tagged at three main sites: Poor Knights Islands (July 1987-August 1989, ;! = 106), Cook Strait (August 1979-June 1984, 7i=599) and Oamaru (March 1988-April 1990, «=918) (Fig. 2). Fish were caught by longline or drop (dahn) line in depths less than 160 m. Only lively fish without everted stomachs or bulging eyes were tagged. Dis- tended swimbladders were vented by means of a hol- low needle inserted through the wall of the body cav- ity. Hapuku were measured (TL), tagged with a loop, dart, or spaghetti tag, and released. Oamaru hapuku were double tagged with both a loop and a dart tag. Loop and spaghetti tags were inserted through the muscle just anterior to the dorsal fin and clipped or tied together, and dart tags were inserted into the muscle directly below the first dorsal-fin ray. Each tag was printed with a serial number, a return ad- dress and notice of a reward. Recapture data and most measurements of recaptured hapuku were provided to the Ministry of Agriculture and Fisheries by the fish- ermen who caught them, but occasionally whole fish were returned for measurement. Von Bertalanffy growth estimates were obtained from the tagging data by using the maximum likeli- hood method and the computer program GROTAG (Francis, 1988). GROTAG also estimates ^^^ and ^p, the mean annual growth of fish of lengths a and (i respectively. The reference lengths a and (i were cho- sen to lie within the range of lengths of tagged hapuku. Francis (1988) showed that these param- eters describe the growth information in tagging data better than the more conventional von Bertalanffy growth parameters L^ and /C do. The expected length increment, AL, for a fish of initial length L^ at lib- erty for time t^T is given by AL = §a -§13 Ll 1 + .AT Also estimated were m and s (the mean and stan- dard deviation of the measurement error), v (the co- efficient of variation of growth variability), andp (the proportion of outliers) (Francis, 1988). Preliminary GROTAG fits suggested that the von Bertalanffy growth model was not appropriate for tagged hapuku; therefore a linear GROTAG function with only one growth parameter ig^^) was fitted to the data. A sea- sonally varying growth function was also tested. Dif- ferences in growth rate between the two sexes in Cook Strait (the sex of recaptured fish was not determined for the other two areas) and among the three tag- ging areas were also investigated. The approach used was to fit a simple GROTAG model with few parameters to the data. The com- plexity of the model was then gradually increased by introducing additional parameters (for example, parameters to allow for seasonal variation in growth). At each stage, new parameter estimates were made and likelihood ratio tests for significant improvement in model fit were carried out as described above for MULTIFAN models. A significance level of 0.05 was used for testing whether there was any gain from introducing additional parameters. Results Otolith ageing Otolith structure The otoliths of the hapuku we ex- amined had an opaque core (dark in transmitted light) and an opaque line that radiated along the dorsoventral axis (Fig. 3). The line was usually split by 1-3 hyaline bands (light in transmitted light). The sulcus was also mostly opaque. The remainder of the otolith consisted of alternating opaque and hyaline bands. In young fish, the hyaline bands were wider than the opaque bands, but difference in width de- clined in older fish. Opaque bands usually consisted of multiple, fine, wavy opaque and hyaline zones (Figs. 3 and 4A). The demarcation between each com- pound opaque band and the adjacent hyaline bands was often indistinct, making the bands difficult to count. The banding pattern was usually clearest in the dorsal half of transverse otolith sections. We used counts of the compound opaque bands as age estimates. Five otolith sections were considered unreadable, and a further four otoliths were lost or damaged dur- ing preparation. Of the remaining 241 otoliths, 78% were scored as poor or moderate (readability 2 or 3); only 22% were good or exceptionally clear (readabil- ity 4 or 5). Validation Otolith marginal state was difficult to assess because of the presence of split opaque rings. There was no apparent seasonal cycle in marginal composition, and sample sizes in most two-month periods were inadequate to provide good estimates of the percentage of the population with opaque mar- gins ( Jan-Feb percent opaque=50.0, n = 14; Mar-Apr 37.5, 8; May-Jun 36.4, 33; Jul-Aug 42.9, 7; Sep-Oct 50.0, 10; Nov-Dec 25.0, 4). The same conclusion was reached for the central data alone («=50). Five OTC-injected hapuku were recaptured after 0.20-2.66 years at liberty (Table 1). All had distinct, bright OTC bands when viewed under UV light (Fig. 4B). There was no apparent relation between OTC 232 Fishery Bulletin 97(2), 1999 Figure 3 Photomicrographs (transmitted white light) of the dorsal halves of transverse sections of hapuku otoliths from (Al an 85-cm female, northern, readability 3, estimated age (mean of R,, R^ j and R^^' H'' years; C = core, S = sulcus, arrows indicate a compound opaque band, circles indicate opaque bands; (Bl an 87-cm female. Cook Strait, readability 4, age 14.7 years; (C) a 106- cm male. Cook Strait, readability 4, age 19.0 years; (D) an 121-cm male, Kaikoura, readability .5, age 45.7 years. Scale bars = 1 mm. Table 1 Details of recaptured hapuku that were tagged and injected — = not determined. Increment width is the amount of po axis. Increment pattern is the type of material (0=opaque margin. An "I" subscript indicates an incomplete band, and a of the otolith margin. Aith o.xy tetracycline (OTC 1. M = male. T = terramycin, E = engemycin, stinjection otolith material deposited along the dorsonuclear otolith , H=hyaline) laid down between the OTC mark (m) and the otolith question mark indicates uncertainty in determining the composition Tag number Length at tagging (cm) Length at recapture (cm) Sex OTC brand Date tagged Date recaptured Years at liberty Increment width (mm) Increment pattern SGO306 89 104 M T 20 Oct 87 19 June 90 2.66 0.61 mO,HOHOH SG0322 8.5 87 M T 8 Jul 88 21 Dec 88 0.45 0.08 mO,H, SG0328 55 55 — E 27 Oct 88 10 Mar 89 0.37 0.16 mO,H, SG0329 73 77 — E 2 Dec 88 12 May 89 0,44 0.05 mHO[? SGO340 68 65 — E 2 Jul 89 14 Sep 89 0.20 0.01 mOiH,'^ brand (terramycin or engemycin) or dosage and the width or intensity of the OTG mark. Four hapuiiu injected with OTC in July-October showed an OTC mark that fell within an opaque band, indicating that opaque bands are formed in the austral winter- spring. The remaining fish was injected in Decem- ber and had an OTC mark at the distal edge of an opaque band, indicating that hyaline band forma- tion begins in spring-summer. The OTC results from all five hapuku are consis- tent with the deposition of one hyaline-opaque band pair on the otolith each year The hapuku that had Francis et al / Age and growth estimates for Polyprion oxygeneios 233 Figure 4 Photomicrographs of the dorsal end of a transverse section from a hapuku otolith illuminated by (A) transmit- ted white light, and iB) reflected W light. The hapuku had been tagged, injected with oxytetracycline, and recaptured after 2.66 years at liberty (see Table 1, tag number SGO306, for further details). Arrowheads indi- cate the position of the inner edge of the oxytetracycline mark, which lies within an opaque band. Outside that band are three hyaline (Hi and two opaque (O) bands. Scale bar = 0.1 mm. been at liberty for the shortest time (SGO340, 0.20 years) had the OTC mark near the margin (postin- jection increment width along the dorsal axis=0.01 mm) (Table 1). Three hapuku with intermediate pe- riods at liberty (0.37-0.45 years) had moderate postinjection otolith increments of 0.05-0.16 mm. SG0328, which was the smallest of the three hapuku at tagging, had the largest increment. SG0322 and SG0328 each had an incomplete opaque band and an incomplete hyaline band outside the OTC mark. SG0329 had a complete hyaline band and possibly an opaque band beginning to form at the margin near the dorsal tip of the otolith. A hapuku that had been at liberty for 2.66 years (SGO306) had a wide postinjection increment (0.61 mm) consisting of a partial opaque band followed by three hyaline and two opaque bands (Fig. 4; Table 1). Age estimates and growth rates R2 j and R., .^ were strongly positively correlated (coefficient of multiple determination [i?'^]=94.6), although they differed markedly for some individuals (Fig. 5A). Some of the larger differences were for otoliths that were judged difficult to count ( readability 2 ), but large differences were also found for clearer otoliths (readability 3-5). R,2 J tended to be higher than R.^ •> for ^^h aged 3-18 years in R.^.^ (mean difference^l.O years) (Fig. 6A). Sample sizes were small for hapuku older than 18, but it appears that the same bias occurred for them. Age estimates made by both readers (Rj and R2 2' were also highly correlated (i?2=92.4) but there were some substantial differences (Fig. 5B). There was no apparent bias between the two sets of estimates for hapuku aged 3-12 in R2 2, but there was a slight ten- dency for R2 2 to exceed Rj over the range 13-18 years (Fig. 6B). Ageing precision was low and variable, both within and between readers, for hapuku aged 3-18 years (Fig. 7). The CV of the age estimates was slightly lower between the two readings of reader 2 (R,, j ver- 234 Fishery Bulletin 97(2), 1999 70 1 . . 60 n = 241 , y/ IT (5 50 o >. I;- 40- a 0) ^ 30. • , T^' ° 20 • <^^' rJfk ■ 10 0 ■ ipE>w> • Readability 3-5 jjP^ D Readability 2 /^ 1 1 . 1 ■ ' /" 60 B / n = 241 y^ 50 ■ / • Age (R,) (years) O O O 10 ^d6C~~ wt^ 0 /^ r 1 1 1 1 1 1 1 0 10 20 30 40 50 60 70 Age (R^ 2) (years) Figure 5 Comparison of hapuku age estimates between (A) the two readings of reader 2 (R,, , and R,, ,,), and (B) reader 1 (Rj) and reader 2 (R, ,). Diagonal lines indicate equality of age estimates. 10 20 30 40 50 Age (R22) (years) Figure 6 Mean hapuku age estimates of (A) reader 2 (R.^ j), and (B) reader 1 (R, ). plotted against the ages determined for the same fish by reader 2 (R., „). The data points are the means and standard errors of the data shown in Figure 5. Diago- nal lines indicate equality of age estimates. sus R22) (mean=18.0%, range=10.1-26.1%) than be- tween readers (Rj versus R22' (niean=21.3'^, range=12.6-31.4%). The CV declined shghtly with increasing age for both the within- and between- reader comparisons. The main differences between counts arose from 1) different band counts near the otoHth core (poten- tial age differcnce=l-3 years), 2) different band counts in an unclear central zone around bands 4-8 (several years), and 3) classification of the composition of the otolith margin as opaque or hyaline (one year). The greatest estimated ages (mean of Rj. R., j and R2 2> were 50.3, 50.7, 59.3, and 63.0 years, but few fish were found to be more than 20 years old (Fig. 5). The youngest hapuku were 3 years old. The von Bertalanffy growth curve fitted to R, data differed little from that fitted to R2 2 data (Fig. 8A; Table 2). Females grew faster than males, but the difference was slight, and the overlap of data points was sub- stantial (Fig. 8B; Table 2). The largest male was 134 cm, the largest female 147 cm (with three females longer than 140 cm). Growth curves fitted separately to the northern and central data were very similar (Table 2). Inspec- tion of the residuals from the central curve again Francis et al,: Age and growth estimates for Polypnon oxygeneios 235 Table 2 Von Bertalanffy growth curve parameters based on hapuku length-at-age and length-frequency data. Standard error estimates for the MULTIFAN parameters are considered u nreliable an d are not shown SE = standard error, Rj = reader 1, R. ., = reading 2 of reader 2, M = male, F = female. Otolith Sample L ±SE <„±SE Sample reading Sex size (cm) K±SE (years) Length-at-age data (otoliths) All ■^2 2 M + F 241 131.7 + 3.2 0.0666 ± 0.0059 -4.63 ± 0.72 All Ri M + F 241 131.0 ± 2.3 0.0785 ± 0.0052 -2.96 ± 0.48 All R22 M 105 129.2 ± 5.4 0.0618 ± 0.0095 -5.75 ± 1.32 All ^2 2 F 110 133.2 ±4.2 0.0698 ± 0.0084 -4.30 ± 1.01 Northern R22 M + F 44 124.0 ±9.8 0.0699 ± 0.0193 -4.51 ± 1.96 Central ^2.2 M + F 172 122.6 ± 5.3 0.0794 ±0.0117 -4.06 ±0.89 Length-frequency data (MULTIFAN) Southern — M + F 1,471 549.2 0.0090 -8.02 suggested that females grew slightly faster than males, although there was strong overlap in the ranges (males: mean residual=-1.8 cm, SD:=8.6, n=90; females: mean=1.6, SD=7.8, n=82). All growth curves had strongly negative values for ^q (Table 2). Length-frequency distributions Hapuku from the four southern trawl surveys gen- erally were less than 85 cm long (Fig. 9), and be- cause 50% maturity occurs at about 85 cm and 88 cm for males and females, respectively (Johnston, 1983), these would have been immature. The small- est hapuku was 41 cm long, but few fish were less than 50 cm. This reflects the pattern seen in the en- tire New Zealand research trawl database which con- tains length measurements for 5841 hapuku: the smallest hapuku was 40 cm, and only 55 were less than 50 cm long. The best-fit MULTIFAN model consisted of con- stant length standard deviation with 11 age classes (see Table 2 for growth parameters). The estimated time elapsed between the theoretical birthday and the first appearance of juvenile hapuku in the length- frequency samples (t^) (based on an estimated mini- mum age at first capture from otolith ageing of 3 years, plus 0.5 years elapsed between the theoreti- cal birthday [1 September] and the sampling date mid- point [1 March] ) was 3.5 years. The age in years since zero length of the youngest age class at the time it first appeared in the length-fi-equency samples (Oj) was es- timated by MULTIFAN to be 11.52 years, producing a t^ estimate of -8.02 years. The standard error estimates provided by MULTIFAN for the growth parameters are not presented in Table 2 because they tended to be unrealistically small (Francis and Francis, 1992). 50 1 40 - c o .« 30 (0 > c (D O it CD O O 20 - 10 5 10 Age (Rj j) (years) 15 Figure 7 Precision of hapuku age estimates of reader 2 (R., ,) and reader 1 (Rjl plotted against the ages determined for the same fish by reader 2 (R,, .,). The data points are the coeffi- cients of variation of the data shown in Figure 4. The growth curve resulting from the best MULTI- FAN fit to the length-frequency data was nearly lin- ear over the range of the data (Fig. 10). The curve was truncated at 14.5 years, which is the maximum age covered by the data (calculated as the sum of the number of age classes detected by MULTIFAN [11] and the estimated age at first recruitment to the samples [3.5 years]). The MULTIFAN curve should not be extrapolated beyond this point because the 236 Fishery Bulletin 97(2), 1999 160 140 120 100 BO 60 40 3 20 I 0 120 100 80 60 40 20 0 5 10 15 20 25 30 35 40 45 50 55 60 65 Age (years) Figure 8 Hapuku length-at-age data (R, ,,), with fitted von Bertalanffy growth cur\'es for (A) all samples and both sexes combined (R, and R, ,1. and (B) all samples combined and males and females separately (R2 2'- See Table 2 for cur\'e equations. A • • • • • • • < < . • ' • • *.M t-' 1 ^ H. . R2^(n = 241) 1 R2.2 R, B . Q — 1 1 1 f 1 1 1 □ D • n -°- = • °ic 0 oo M^l^ • *ad 3 CMB^O Wl •fi • ^\ * ; r* • Males(n=105) D Females (n= 110) -'f Males 1 1 1 — Females length-frequency data represent only small, imma- ture hapuku. Over the age range 0-12 years, the MULTIFAN growth curve was very similar to growth curves based on otolith ages (Fig. 10). Tag recapture One tagged hapuku was recaptured, released with a new tag, and recaptured a second time (Johnston. 19921. It grew from 73 cm to 83 cm during its first 800 days at liberty and from 83 cm to 96 cm during its second 829 days at liberty. This fish was entered into the growth data set three times: once for each of the two periods at liberty and once for the combined period at liberty. All other hapuku used in the growth analysis were recaptured only once. One hundred and forty four recaptured hapuku had sufficient information (lengths at tagging and recap- ture, days at liberty) to be used in the growth analy- sis. Most of these hapuku were between 50 and 90 cm long at tagging (Fig. 11) and would have been immature. The GROTAG reference lengths a and /3 were set at 55 cm and 85 cm respectively. Periods at liberty ranged from 9 to 3708 days (0-10.2 years), but most fish were at liberty for less than four years (Fig. 12). Initial GROTAG model fits identified six fish as outliers (fish having absolute standardized residu- als greater than 3.0). Five of these fish had negative growth increments of 3-27 cm, and one had a posi- tive increment of 24 cm after 347 days at liberty. These outliers probably resulted from measurement Francis et al.: Age and growth estimates for Polyprion oxygeneios 237 40n TAN9301 n = 434 30 40 50 60 70 80 90 100 110 120 130 Total length (cm) Figure 9 Hapuku length-frequency distributions (both sexes combined) ob- tained from four Tangaroa trawl surveys off Southland in Febru- ary-March 1993-1996. n = sample size. errors and were deleted from the data set, leaving 138 fish. A simple linear model, with a single growth-rate parameter, produced an estimated gi-owth rate of 4.26 cm/yr (Table 3, model 1). A model with two growth- rate parameters produced no improvement in fit (Table 3, model 2). The addition of parameters for outlier contamination, measurement bias, and sea- sonal growth did not significantly improve the log- likelihood (Table 3, models 3-5), nor did subdivision of the data into three geographical areas (Table 3, model 6). For the Cook Strait data subset, inclusion of separate growth parameters for male and female hapuku did not significantly improve the model fit (Table 4). However, sample sizes were small, statis- tical power was low, and only large differences could have been detected. Therefore the best GROTAG model (Table 3; model 1) consisted of simple linear growth over the length range 55-85 cm. It had no apparent pattern or trend in the residuals. GROTAG was then run in simula- tion mode (Francis, 1988) by using parameter esti- mates from model 1 to determine the accuracy and precision of the growth-rate estimate. The mean an- nual growth rate was estimated to be 4.25 cm/yr with a standard error of 0.017 cm/yr. 238 Fishery Bulletin 97(2), 1999 125 1 100 - "e ^ 75 jC B c « 50- 25 ■ 0 0 -- i£? Otoliths, txjth sexes • Otoliths, males Otoliths, females LF, both sexes • l^cDougall, males 0 McDougall, females 0 5 10 15 20 Age (years) Figure 10 Comparison of growth curves derived from length-at-age data (R2,,. otoliths), and MULTIFAN analysis of length-frequency data (LF). Also shown are the mean lengths-at-age reported by McDougall 1197.5) for hapuku collected in Cook Strait. Table 3 Parameter estimates from GROTAG growth models for hapuk J from all tagging sites Ui = 138 . Seasonal phase is given in years since 1 January. SD = standard deviation; — = parameter not estimated . Models 1-5 are for Poor Knights I slands. Cook Strait, | and Oamaru, data combined; model 6 estimated g^j separately for the th ree areas. Areas Parameter All areas combined separate 1 2 3 4 5 6 Log likelihood 2X 697.80 697.42 697.80 697.80 694.30 695.85 Mean growth rates g-^ (cm/yr) 4.26 4.01 4.24 4.25 4.25 4.1-5.0 Ss!, (cm/yr) — 4. .38 — — — — Growth variability V 0.38 0.39 0.39 0.39 0.39 0.39 •SD measurement error s (cm) 1.55 l.,54 1.55 1.55 1.51 1.51 Measurement bias m ( cm ) — — — 0.05 — — Outlier contamination P — — 0.00 — — — Amplitude u — — — — 0..53 — Phase IV (yri — — — — 0.15 — Discussion Hapuku otoliths frequently had split opaque bands and were difficult to age. This led to reduced count- ing precision and, for some age groups, to a small between- and within-reader ageing bias. Our use of marginal increment analysis to validate hapuku otolith ageing was inconclusive. We attribute this to the compound nature of the opaque bands and to the resulting difficulty in determining whether the ma- terial being deposited at the margin was opaque or hyaline. The results of the OTC injection experiment were consistent with the deposition of one opaque and one hyaline band per year. This does not constitute full validation of the otolith ageing technique because the Francis et al.: Age and growth estimates for Polyphon oxygeneios 239 1 - in a 8 a. 0) E 2 3 0 10 8 6 4 2 0 Poor Knights n=16 Cook Strait n=57 Oamaru n = 65 40 60 70 80 Total length (cm) 100 Figure 11 Length-frequency distributions (at tagging) of tagged and recaptured hapuku used in ttie GROTAG analysis of growth rates, n = sample experiment included only five hapuku, only one of which was at liberty for more than one year Further- more, all five fish would have been immature or just maturing at the time of injection. Further validation is required, especially for older, mature hapuku. An independently derived von Bertalanffy growth curve (based on MULTIFAN analysis of length-fre- quency data) agreed well with the otolith-based length-at-age curves over the age range 0-12 years but indicated faster growth rates for hapuku aged 12-14.5 years (Fig. 10; Table 5). Our previous expe- rience has shown that MULTIFAN gi'owth curves are generally reliable for younger age classes, which of- ten exhibit discernible length modes, but are not re- liable for older age classes, which usually lack any modal length structure. This is because MULTIFAN underestimates the number of older age classes present in the data and consequently overestimates the length-at-age of the older age classes (Francis and Francis, 1992; Francis, 1997 ). Therefore, the dis- crepancy between the MULTIFAN and length-at-age curves for the older age classes in Fig. 10 probably Table 4 Parameter estimates from GROTAG growth models for hapuku from Cook Strait (/i=43). SD = Stan dard devia- tion; — = parameter not estimated Sexes separate Both Parameter sexes M F Log likelihood 2X 240.44 240.26 Mean growth rates g^^ (cm/yr) 4.53 4.44 4.64 g,- Icm/yrl — — Growth variability I' 0.27 0.27 SD measurement error s (cm) 1.94 1.99 arises from bias in the former This interpretation is supported by the near-linearity of the MULTIFAN growth curve and the implausible MULTIFAN growth parameters, especially L^ (Table 2). 240 Fishery Bulletin 97(2), 1999 O 60 50 - 40 30 ■ E o_ c « E 0) o .E 20 % 2 10 -10 A Poor Knights (n= 16) ■ Cook Strait (n= 57) o Oamaru (n = 65) 1,000 2,000 Days at liberty 3,000 4,000 Figure 12 Relationship between growth increment and period at liberty for tagged hapuku. n = sample size. Table S Comparison of hapuku annual growth increments at two reference lengths (5.5 cm and 85 cmi based on length-at- age (R^., ), length-frequency, and tag-recapture data. Annual growth increment (cm) Total length Tag- (cmi Length-at-age Length-frequency recapture The best GROTAG model fit e.stimated the mean annual growth rate to be 4.25 cm/yrover the length range 55-85 cm. That estimate falls between the length-at-age estimates of annual growth for 55-cm and 85-cm hapuku (Table 5). A nonlinear growth model was not significantly better than the linear model, probably because of the narrow length range of the tag-recapture data, and the small sample size (n = 138). We conclude that the growth curves and growth rate estimates derived from length-frequency and tag-recapture data are consistent with those from otolith-based, length-at-age data for the length and age ranges over which the data overlap. In conjunc- tion with the results of the OTC injection experiment, this agreement among growth-rate estimates pro- vides strong support for our hypothesis that otolith band pairs are deposited annually in hapuku. Our length-at-age growth curve for male hapuku is similar to that of McDougall (1975), who made band counts in broken and burnt otoliths from hapuku collected in Cook Strait (Fig. 10). However, McDougall reported a slightly faster growth rate for females (Fig. 10). It is not possible to determine whether the difference in female growth rates is real, or a result of McDougall's small sainple size (n=72). McDougall's largest hapuku was 120 cm and his old- est fish was 26 years. A study of hapuku growth at Juan Fernandez Is- land, based on scale annuli, reported much faster growth rates and a maximum age of only 12 years (Pavez and Oyarzun. 1985). Scale rings are typically crowded to the point of being unresolvable in hapuku older than 8 years (McDougall, 1975); it is likely therefore that the ages of Pavez and Oyarziin's larger fish were underestimated. Lack of precision and reader bias mean that otolith ageing is unlikely to be useful when accurate hapuku ages are required (e.g. in the estimation of year-class strength from an age-frequency distribution ). How- ever, the precision and accuracy problems experi- enced in this study had little effect on the shape of growth curves fitted to length-at-age data, and we believe that the growth parameters reported here are relatively robust. The large negative t^ values for all growth curves indicate a lack of fit of the von Bertalanffy model to the data for small hapuku. This may result from one or more of the following: 1 ) lack offish less than three Francis et al.: Age and growth estimates for Polyphon oxygeneios 241 years old in our data; 2) underestimation of the ages of hapuku because of failure to recognize and count all the annual bands near the core; or 3) different growth rates of pelagic juveniles, and demersal ju- veniles and adults. Interestingly, Pavez and Oyar- ziin's (1985) scaled-based growth curves had t^ val- ues close to zero, and estimated ages at the length of settlement similar to ours (i.e. 38 cm at age 3 and 49 cm at age 4). Scale bands may therefore help to re- solve problems experienced in interpreting otolith cores when ageing small hapuku. Small juvenile hapuku pass through a pelagic stage in surface waters, usually associated with flotsam, and are rarely caught (Cormack, 1986; Michael, 1988; Roberts, 1996). At the end of the pelagic stage, they become demersal in depths of 50-600 m. Bottom trawl length-frequency data (Fig. 9; NIWA") and bottom set-net and line records (Johnston, 1983; Roberts, 1989) indicate that most hapuku switch from a pelagic to demersal life style at around 50 cm, although hapuku as small as 40 cm have been caught by bottom trawl. A 35-cm hapuku reported caught in Wellington Harbour by Hector ( 1888) may have been a pelagic juvenile, but no details were given of habitat or fishing method. A small number of pelagic juveniles up to 67 cm long have been caught on surface longlines over 2000 m or more of water (Roberts, 1996; Scientific Observer Database'^). These large pelagic juveniles may have become "stranded" in the pelagic environment after being carried over deep water by oceanic currents. Presum- ably they remain in surface waters until they en- counter shallower depths, at which time they settle to the seabed. Our best estimate of the age at which hapuku settle to a demersal habitat is 3-4 years, but our interpre- tation of the banding pattern near the otolith core was uncertain and somewhat subjective. Counts of possible daily increments in otolith sections also sug- gest settlement at about 3 years (Roberts, 1996), and a similar age range has been suggested by other stud- ies (McDougall, 1975; Johnston, 1983). Polyprion americanus apparently settles at lengths greater than 45-55 cm and ages greater than 1-7 years (Sedberry et al., 1996). The age at recruitment of hapuku to commercial trawl catches is likely to be the same as the age at settlement to the seabed. Although our research ves- sel samples were caught with a small mesh (60-mm) codend, hapuku longer than 50 cm are also retained by the larger mesh sizes used by commercial vessels (typically 100-125 mm) (Hurst and Bagley 1997). Previous studies have suggested that female hapuku grow larger and faster than male hapuku (McDougall, 1975; Roberts, 1986; 1989). Our results support those conclusions but also indicate that the differences in both maximum length and growth rate are small and, for most purposes, trivial. However, our ability to detect growth rate differences between the two sexes, and between sample sites, was lim- ited by the low precision of our age estimates and the small sample sizes available for the GROTAG analysis. Hapuku mature at about 85 and 88 cm, males and females, respectively (Johnston, 1983). Based on the length-at-age growth curves given in Table 2, age at maturity is estimated to be 10-13 years for both sexes. The largest hapuku in our study was 147 cm TL, but they are known to reach 178 cm TL and 76 kg in weight (Roberts^). The longevity of hapuku cannot be precisely determined because of difficulties we experienced in estimating ages from otoliths, and because our samples mostly came from populations of hapuku that had been exploited for many years. Our study does suggest that hapuku live at least 50 years, possibly in excess of 60 years. Acknowledgments We thank Alex Johnston (formerly Ministry of Agri- culture and Fisheries) and Clive Roberts (Museum of New Zealand-Te Papa Tongarewa, formerly Victoria University of Wellington) for providing hapuku otoliths. We are also grateful to Alan Coakley, John Holdsworth, Alex Johnston, and Peter Saul ( all formerly Ministry of Agriculture and Fisheries) for conducting the tagging programs. Their combined efforts provided the basis for much of this study. C. D. Roberts and J. A. Colman made useful comments on an earlier draft. This research was carried out by NIWA under contract to the New Zealand Ministry of Fisheries (contract no. PIAGOl). Literature cited Annala, J. H., and K. J. Sullivan (eds). 1997. Report from the Fishery Assessment Plenary, May 1997: stock assessments and yield estimates. Ministry of Fisheries, Wellington, 381 p. ^ NIWA (National Institute of Water and Atmospheric Research I. ^ P. O. Box 14901. Welhngton, New Zealand. Unpubl. data. "^ Scientific Observer Database, NIWA, Wellington, New Zealand. Unpubl. data. ^ Roberts, C. D. 1997. Museum of New Zealand-Te Papa Tongarewa, P. O. Box 467, Wellington, New Zealand. Personal commun. 242 Fishery Bulletin 97(2), 1999 Bagley, N. W., and R. J. Hurst. 1995. Trawl sui-\'ey of middle depth and inshore bottom species off Southland, February-March 1994 (TAN94021. N. Z. Fish. Data Rep. 57, 51 p. 1996a. Trawl survey of middle depth and inshore bottom species off Southland, February-March 1995 (TAN9502I. N. Z. Fish. Data Rep. 73, 48 p. 1996b. Trawl survey of middle depth and inshore bottom species off Southland. February-March 1996 (TAN9604I. N. Z. Fi.sh. Data Rep. 77, 51 p. Campana, S. E., M. C. Annand, and J. I. McMillan. 1995. Graphical and statistical methods for determining the consistency of age determinations. Trans. Am. Fish. Soc. 124:131-138. Cormack, S. 1986. Observations on warehou. Catch 13(11):12. Fournier, D. A., J. R. Sibert, J. Majkowski, and J. Hampton. 1990. MULTIFAN a likelihood-based method for estimat- ing growth parameters and age composition from multiple length frequency data sets illustrated using data for south- ern bluefin tuna iThunnus maccoyii). Can. J. Fish. Aquat. Sci. 47:301-317. Francis, M. P. 1997. Spatial and temporal variation in the growth rate of elephantfish iCallorhinchus milii). N. Z. J. Mar Fresh- water Res. 31:9-23. Francis, M. P., and R. I. C. C. Francis. 1992. Growth rate estimates for New Zealand rig iMustelus lenticulatus). Aust. J. Mar Freshwater Res. 43:1157- 1176. Francis, R. I. C. C. 1988. Maximum likelihood estimation of growth and growth variability from tagging data. N. Z. J. Mar. Freshwater Res. 22:42-51. Hector, J. 1888 On a small-sized specimen of the hapuka, Hectoria (Oligorus)gigas.Caste\neau. caught in Wellington Harbour. Trans. Proc. N. Z. Inst. 20:446-447. Hurst, R. J., and N. W. Bagley. 1994. Trawl survey of middle depth and inshore bottom -species off Southland, February-March 1993 (TAN9301). N. Z. Fish. Data Rep. 52, .58 p. 1997. Trends in Southland trawl surveys of inshore and middle depth species, 1993-96. N. Z. Fish. Tech. Rep. 50, 67 p. Johnston, A. D. 1983. The southern Cook Strait groper fishery. Min.Agric. Fish. Fish. Tech. Rep. 159, 33 p. 1992. Groper tag recovery exceeds 10 years. N. Z. Profes- sional Fisherman 6(71:59. Kailola, P. J., M. J. Williams, P. C. Stewart, R. E. Reichclt, A. McNee, and C. Grieve. 1993. Australian fisheries resources. Bureau of Resource Sciences, Department of Primary Industries and Energy, and Fisheries Research and Development Corporation, Canberra, 422 p. McDougall, C. R. 1975. Age and growth of Polyprion oxygeneios in Cook Strait. B.Sc. (Honours) diss., Victoria Univ. Wellington, Wellington, 45 p. Michael, K. 1988. Unusual longline catches and observations. Catch 1,5(11):16. Paul, L. J. , and N. M. Davies. 1988. Groper N. Z. Fish. Assess. Res. Doc. 88/15, 27 p. Pavez, P., and M. E. Oyarziin. 1985. Determinacion de eficiencia relativa de espineles y parametros de crecimiento del bacalao de Juan Fernandez (Polyprion oxygeneios Bloch y Schneider, 1801 1, en las islas Robinson Crusoe y Santa Clara. In P. Arana ( ed.l, Investigaciones marinas en e archipielago de Juan Fernan- dez, p. 323-340. Universidad Catblica de Valparaiso, Valparaiso. Pizzaro, L. G., and E. R. Yanez. 1985. Estimacion de la edad de primera captura y de la mortalidad por pesca optimas del bacalao de Juan Fernandez iPolyprion oxygeneios Bloch y Schneider, 1801 ), a traves del analisis del rendimiento por recluta. In P. Arana ( ed.), Investigaciones marinas en e archipielago de Juan Fernandez, p. 341-345. Universidad Catolica de Valparaiso, Valparaiso. Roberts, C. D. 1986. Systematics of the percomorph fish genus Polyprion Oken, 1817. Ph.D. diss., Victoria Univ Wellington, Wel- lington, 283 p. 1989, Reproductive mode in the percomorph fish genus Polyprion Oken. J. Fish Biol. 34:1-9. 1996. Hapuku and bass: the mystery of the missingjuveniles. Seafood N. Z. 4(11:17-21. Sedberry, G. R., J. L. Carlin, R. W. Chapman, and B. Eleby. 1996. Population structure in the pan-oceanic wreckfish, Polyprion americanus (Teleostei: Polyprionidael, as indicated by mtDNA variation. .J. Fish Biol. 49 (suppl. Al:318-329. Sedberry, G. R., G. F. Ulrich, and A. J. Applcgate. 1994. Development and status of the fishery for wreckfish, (Polyprion americanus) in the southeastern United States. Proc. Gulf and Caribbean Fish. Inst. 43:168-192. 243 Abstract.— This study examined the relative rates of billfish bycatch and target species catch by areas ( r\ 2°, and 5 latitude and longitude) and months m the catch data reported in mandatoi-y log books kept by U.S. pelagic-longline fishermen in order to identify potential time-area strata that could reduce bill- fish bycatch. The 1986-91 mean per- centages identified month-area strata with high percentages of sailfish and marlin bycatch and marlin only bycatch. The analyses indicated that the elimi- nation of effort in cells selected accord- ing to percentages of billfish in the catch could have reduced the 1986-91 billfish bycatch by 509; and the target species from 13.9 to 19.2%, depending on the spatial resolution employed. The corresponding analysis of marlin only indicated a 509f reduction in marlin bycatch could have been attained and a 16.4-20.7'? reduction in the target species catch. The time-area closures identified in the 1986-91 logbook data were applied to the data for 1992-95 and provided a test of the spatial and temporal stabilities of these results. For the evaluation of sailfish and marlin combined, the reductions in both bill- fish bycatch and target species catches averaged less than the predicted val- ues, but in all cases billfish were selec- tively protected. For the evaluation of marlin only, the reduction of sailfish bycatch was less than the predicted amount and the reduction of the target species was slightly greater than the predicted value. The agreement be- tween the predicted level of protection for billfish or marlin and the mean value for the 1992-95 test period in- creased with increasing size of the grid. At the 5'cell size, the mean reduction was 22.8% for the targeted species and 48.6% for marlin (compared with pre- dicted values of 20.7 and 50% respec- tively). These results suggest that time and area restrictions on fishing could significantly reduce the bycatch of billfishes in the pelagic-longline fish- eries without equivalent reductions in the catch of target species. An analysis of the possible utility of time-area closures to minimize billfish bycatch by U.S. pelagic longlines C. Phillip Goodyear 415 Ridgewood Road Key Biscayne, Florida 33149 E-mail address phiLgoodyear g email msn com "Manuscript accepted 22 June 1998. Fish. Bull. 97:243-255 (1999). The U.S. Fisheries Management Plan for Atlantic Billfishes ( sailfish, blue and white marlin, and spear- fish) reserves billfish stocks for recreational use; their commercial hai^vest by U.S. fishermen has been prohibited since the plan became effective in 1988 (SAFMC, 1988). The most recent stock assessment for billfishes found that, over the Atlantic as a whole, blue and white marlin stocks are seriously over- fished (Anonymous, 1996). Their stocks were at about 21 and 24%, respectively, of the biomass needed to support maximum sustainable yield ( MSY) at the beginning of 1996 (Anonymous, 1996). Fishing-in- duced mortality was estimated to be 3.19 times higher than that which would produce MSY for blue mar- lin, and 1.88 times higher than that which would produce MSY for white marlin. Most of the fishing mortal- ity on these species is the result of bycatch in the pelagic-longline fish- eries that target other species, par- ticularly tuna and swordfish. Many authors have promoted the concept of marine reserves (also called marine protected areas or MPAs) as management tools to enhance conservation of fishery resources (Bohnsack, 1994; Shackell and Wilson, 1995; Hutchings, 1995, 1996; Allison et al., 1998; Lauck et al., 1998). Closed areas have been used in many parts of the world to control bycatch mortality (Alverson et al., 1994). Hutchings { 1996 ) noted that MPAs might have considerable merit in reducing bycatch when as- sessed against the effectiveness of other forms of regulatory control such as bycatch limitations and catch quotas. Often, however, the concept of marine reserves is re- stricted to closures of particular habi- tats or portions of habitats. Kench- ington (1990) noted that such marine reserves would be of little use for species such as billfish that have both pelagic larvae and highly pelagic adults. Among other options for reducing bycatch, Alverson et al. (1994) observed that time-area control of fishing activity offers an opportunity to reduce unwanted bycatch. The main objective of employing time— area strategies is to take advantage of variation in the degree of co-occurrence between tar- get and bycatch species (Murawski, 1992). The effectiveness of such controls obviously depends on the degree of overlap between the bycatch and the targeted species (Adlerstein and Trumble, 1992). Cramer ( 1996 ) used logbook data on U.S. large pelagic fish (data that have been mandatory since October 1986) in order to evaluate the spatial distribution of the catch of undersized swordfish by the long- line fishery in strata of 1° latitude and longitude by season and area. Her study evaluated the implica- tions of removing time-area strata with high swordfish discard rates on the total catch of the longline fish- ery. The present study uses the same data set to evaluate the degree of 244 Fishery Bulletin 97(2), 1999 overlap between billfish bycatch and target species catch for the U.S. pelagic longline fishery and develops a method for selecting candidate time— area cells whose closure would selectively enhance billfish survival and minimize impacts on the target species catch. culating the percent reduction in the total billfish and marlin bycatch that would have occurred each year if the time-area cells identified in the 1986-91 data had been closed. This analysis assumed that the displaced effort would not shift elsewhere. Methods Results The longline catch data were inspected 1) to determine the species composition of the catch to identify the set of species which, taken together constitute the primary targets of the fishery, and 2 ) to identify possible trends in the reporting of billfish bycatch. The reported bycatch of billfish declined markedly in 1992 compared with earlier years. Consequently, the data were partitioned into two periods: 1986-91 and 1992-95. Data fi-om the first period were used to characterize the spatial and temporal distributions of the catch. Data from the sec- ond period were used to test temporal stability of the results of analyses of the first period. The billfish bycatch and the combined billfish (sail- fish, spearfish, and blue and white marlin) bycatch as well as the target species catch for each longline set of each reported trip were summed into cells of 1°, 2°, and 5° latitude and longitude and month over the period 1986-91. The analysis was also repeated and only blue and white marlin bycatch was consid- ered along with the target catch. Separate cumula- tive frequency distributions of the percentage of bill- fish or marlin bycatch in the combined billfish and target species catches by number by cell were con- structed for each level of spatial resolution. These frequency distributions were used to estimate the percent reduction in billfish bycatch and target spe- cies catch that would have occurred if the effort in time-area cells with billfish or marlin bycatch rates above an arbitrary threshold had been completely eliminated during the 1986-91 fishery. The distri- bution of the minimum number of cells producing any specific reduction in billfish or marlin catches with the least impact on target species catch could then be determined from these data. As an example, the distribution of cells that, if they had been eliminated from the 1986-91 U.S. fishery, would have reduced the billfish and marlin bycatches by SO'/f , were plotted for examination. Although the reporting rate declined after 1991, the temporal and spatial relative distributions of billfish in the total longline catch would be unaffected, so long as the change in the reporting rate was random. Conse- quently, the temporal stability of the reduction in billfish and marlin bycatch for the time-area cells identified with the 1986-91 data was evaluated by using data from 1992 to 1995. This was done by cal- The catch by species in the data file from U.S. pe- lagic longline logbooks from 1986 through 1995 is presented in Table 1. Inspection of these data indi- cates that the predominant target species of this fish- ery included swordfish, yellowfin tuna, bigeye tuna, albacore, and dolphin fish. Together they constituted 90% of the nonbillfish harvest that was reported in these logbooks. For this analysis, billfish bycatch is the combined catch of sailfish, spearfish, and blue and white marlin, and marlin bycatch is the com- bined catch of blue and white marlin. Inspection of the data in Table 1 indicates a sharp drop in the numbers of billfish caught after 1991. This is possi- bly a consequence of a change in reporting trends following the U.S. billfish management plan's prohi- bition of the sale of billfish, or the restrictions on landing undersized swordfish imposed in mid-1991, rather than a consequence of a true reduction in the bycatches of billfish by the U.S. pelagic-longline fleet. Although no obsei-ver data are available before 1992, this conclusion is supported by data collected by ob- servers on longline vessels that indicate billfish bycatch rates after 1992 were much higher than those reported in the logbooks ( Cramer M. For this reason, the longline logbook data were grouped into two pe- riods: 1986-91 and 1992-95. Data from 1986 through 1991 were used to characterize the distribution of billfish and target species catches, and data from 1992 through 1995 were used to test the spatial and temporal stability of the patterns observed in the 1986-91 data. The percentage of billfish in the combined billfish bycatch and target species catch for each month-area cell in strata of 1°, 2" , and 5° was estimated, and the results were sorted to obtain cumulative frequency distributions of the billfish bycatch percentages for each spatial resolution. At the same time, the cumu- lative catches of billfish and target species were com- piled for each observation in the cumulative fre- quency distributions of percentage of billfish in the catch in each time-area cell. The results of this pro- cedure are presented in Figure 1 and Table 2. Note Cramer. J. 1996. Pelagic longline billfi.sh bycatch. Inter- national Commission for the Conservation of Atlantic Tunas (ICCATi working document SCRS/96/97. Goodyear: Analysis of time-area closures to minimize billfish bycatch 245 Table 1 Catch in number (not including billfish discards) by species reported in the logbook data for U.S. pelagic long] ines for 1986-95. Species 1986 1987 1988 1989 1990 1991 1992 1993 1994 1995 Total Swordfish 16.510 113,665 161,980 171,344 129,147 86,586 67,004 64.724 63,913 72,520 947,393 Bluefin tuna 58 484 422 717 599 540 1010 342 335 266 4773 Albacore 1181 2085 2984 3986 8220 9873 14.137 13,636 18.795 21.810 96,707 Bigeye tuna 3162 15,919 15,249 19.764 16.947 18,250 15,239 21.446 22.859 25.363 174,198 Yellowfin tuna 9241 59,665 67.569 83.844 68.711 70,433 94.510 68.360 73.337 85,135 680,805 Blackfm tuna 64 560 808 1450 1090 1525 3178 2797 3096 2363 16,931 Skipjack tuna 0 0 0 0 0 1384 2165 1374 232 279 5434 Bonito 0 0 0 0 0 0 0 238 1030 497 1765 Other tuna 0 0 0 0 0 0 1695 613 337 451 3096 Blue marlin 115 815 387 65 82 55 3 64 62 23 1671 Sailfish 4 55 147 53 66 22 5 34 56 87 529 Spearfish 4 52 47 2 15 2 8 30 19 14 193 White marlin 217 998 468 97 48 62 13 36 27 20 1986 Greater amberjack 0 0 0 0 0 0 112 836 1278 234 2460 Banded rudderfish 0 0 0 0 0 0 3 13 29 86 131 Dolphin fish 972 7861 10,768 29,899 31.899 32.576 25,.351 26.142 29.617 72,321 267,406 King mackerel 1 11 35 119 11 67 118 4602 7176 1929 14,069 Wahoo 40 1346 2629 3157 3560 3736 6506 4393 3442 5477 34,286 Oilfish 0 0 0 0 0 0 0 37 5073 5102 10,212 Blue shark 944 1482 581 980 633 753 2050 1513 864 373 10,173 Mako shark 486 4055 4877 6946 4380 3475 414 338 371 187 25,529 Shortfin mako shark 0 0 0 0 0 0 3944 3738 3982 3747 15,411 Oceanic whitetip shark 0 0 0 0 0 0 319 226 166 205 916 Porbeagle shark 0 0 0 0 0 0 41 675 1,591 509 2816 Bigeye thresher shark 0 0 0 0 0 0 200 304 626 472 1602 Other pelagic shark 0 0 0 0 0 0 0 0 0 8 8 Thresher shark 32 674 632 488 280 194 113 139 95 181 2828 Dusky shark 0 0 0 0 0 0 3365 7305 2826 2830 16,326 Bignose shark 0 0 0 0 0 0 70 254 124 123 571 Blacktip shark 0 0 0 0 0 0 3527 5965 6042 3954 19,488 Hammerhead shark 57 129 535 979 500 262 360 562 508 349 4241 Scalloped hammerhead shark 0 0 0 0 0 0 292 267 646 375 1580 Smooth hammerhead sh ark 0 0 0 0 0 0 627 452 484 231 1794 Night shark 0 0 0 0 0 0 478 455 379 421 1733 Silky shark 0 0 0 0 0 0 1685 1244 2134 2765 7828 Spinner shark 0 0 0 0 0 0 925 255 234 209 1623 Tiger shark 27 104 1051 183 98 108 165 173 477 202 2588 Other sharks 842 5740 4380 5424 9750 4620 3242 4242 1092 1376 40,708 White shark 20 137 1723 1222 537 95 104 163 16 2 4019 Sandbar shark 0 0 0 0 0 0 0 0 2020 14,192 16,212 Other 0 0 0 0 0 0 0 0 90 2515 2605 that the initial slope of the relation between the bill- fish percent reduction and target species percent re- duction is much lower than 1.0 for each of the spa- tial resolutions examined. This finding suggests that the temporal and spatial distributions of billfish bycatch and target species are somewhat different and that it should be possible to selectively reduce the longline billfish bycatch in relation to target spe- cies catch. The data from Table 2 indicate that a 10% reduction in the 1986-91 billfish bycatch could have 246 Fishery Bulletin 97(2), 1999 Table 2 Estimated reduction in billfish bycatch and target spec es catch bv elimination of time-area c« lis where the biUfi ■ih percentage of catch is greater than or equal to the specified threshold value based on 1986 -1991 catch rates m the U.S. pt lagic- ongline fishery. These estimates assume that the displaced effort will not be redirected elsewhere. Billfish 1 ° cells 2° cells 5°cells Target Target Target reduction (%) Threshold (%) reduction (%) Threshold (9c ) reduction (9c) Threshold ( 9c) reduction (9c ) 5 18.46 0.34 14.23 0.49 9.64 0.94 10 14.32 1.01 9.99 1.51 8.77 2.33 15 12.06 1.90 9.44 2.73 7.94 3.88 20 10.07 2.97 8.52 4.08 7.58 5.52 25 8.81 4.25 7.17 5.62 7.34 7.21 30 7.52 5.74 5.86 7.54 6.10 9.19 35 6.78 7.45 5.47 9.70 5.70 11. .38 40 6.16 9.37 4.87 12.13 5.52 13.68 45 5.49 11.49 4.61 14.77 4.86 16.29 50 4.95 13.87 4.32 17.57 4.36 19.17 55 4.37 16.59 4.10 20.53 4.21 22.28 60 4.03 19.59 3.79 23.72 3.95 25.51 65 3.61 22.85 3.51 27.16 3.57 29.11 70 3.30 26.53 3.21 30.92 3.11 33.13 75 2.94 30.62 2.95 35.01 2.76 37.67 80 2.56 35.30 2.48 39.73 2.46 42.92 85 2.17 40.73 2.19 45.21 2.07 48.85 90 1.74 47.38 1.70 52.09 1.73 55.91 95 1.20 56.08 1.15 61.54 1.26 65.08 been attained with only a 1 to 2.3% reduction in the catch of target species, depending on the spatial resolution employed. This reduction would have been achieved by eliminating the effort in months and areas where the reported billfish bycatch exceeded the threshold percent- age of billfish in the combined catch in Table 2. Similarly, a 509c reduction in the 1986-91 bill- fish bycatch would have been achieved by elimi- nating the effort in cells where the reported bill- fish bycatch exceeded the table values with cor- responding reductions of 13.9 to 19.2% in the harvest of target species. A 95% reduction in the 1986-91 billfish bycatch would have been achieved by eliminating all effort in cells with reported catches of more than 1.2, 1.15, or 1.26% billfish for 1°, 2°, or 5° cells, and there would have been a concomitant reduction in target species catch of 56.1, 61.5, or 65.1%, respectively. The same type of analysis was conducted with only blue and white marlin as the billfish spe- cies of concern. The results are presented in Figure 2 and Table 3. As with billfish combined, the initial slope of the relation between the per- cent catch of marlin bycatch and of target species is much less than 1.0. This result also indicates that the combined marlin bvcatch can be reduced by the (/) o o Q. (D 0 10 20 30 AO 50 60 70 60 90 Billfish percentage reduction Figure 1 Percent reduction in target species and marlin bycatch associ- ated with the removal of effort in time-area cells above the indi- cated threshold percentage of all billfish in the catch for cell sizes of V, 2", and 5= of latitude and longitude. elimination of select time-area strata and that the catch of target species would not be proportionately affected. The data from Table 3 indicate that a 10% Goodyear: Analysis of time-area closures to minimize billfish bycatch 247 Table 3 Estimated reduction in blue a nd w lite marlin bycatch and target species catch by elimination of time- area cells where the | marlin percentage of catch is greater th an or equal to the specified threshold value based on 1986-91 ca tch rates in the U.S. pelagic-longline fishery. These estimates assume that the displaced effort will not be redirected elsewhere Marlin 1 'cell s 2° cells 5'cells Target Target Target reduction (%) Threshold (%) reduction (%) Threshold (%) reduction {%) Threshold (%) reduction (%) 5 18.15 0.38 11.54 0.53 7.32 1.11 10 10.16 1.14 7.06 1.71 6.22 2.79 15 7.71 2.34 5.19 3.50 5.49 4.76 20 6.17 3.79 4.79 5.37 5.21 6.33 25 5.77 5.10 4.24 7.44 4.83 8.62 30 4.97 6.80 4.02 8.99 4.65 10.81 35 4.54 8.57 3.87 11.59 4.16 13.63 40 4.15 10.95 3.64 14.14 4.02 15.70 45 3.78 13.60 3.44 16.55 3.86 17.72 50 3.46 16.37 3.23 19.55 3.58 20.74 55 3.24 19.50 3.12 22.44 3.12 23.66 60 3.00 22.03 2.87 25.95 2.84 27.44 65 2.76 25.04 2.67 29.47 2.68 30.54 70 2.39 28.50 2.39 32.87 2.47 34.89 75 2.14 32.94 2.10 36.53 2.20 38.76 80 1.86 37.90 1.85 41.40 2.01 44.08 85 1.56 43.97 1.67 46.43 1.78 49.78 90 1.21 50.49 1.23 53.72 1.49 56.75 95 0.73 59.39 0.84 63.06 1.07 66.45 reduction in the 1986-91 marlin bycatch could have been attained with only a 1.1 to 2.8% re- duction in the catch of target species, depend- ing on the spatial resolution employed. This would result by eliminating the effort in months and areas where the reported billfish bycatch exceeded the threshold percentage of billfish in the combined catch indicated in Table 3. Simi- larly, a 50% reduction in the 1986-91 marlin bycatch would have been obtained by eliminat- ing the effort in cells where the reported bill- fish bycatch exceeded the table values with cor- responding reductions of 16.4 to 20.7% in the harvest of target species. A 95% reduction in the 1986-91 marlin bycatch would have been achieved by eliminating all effort in cells that reported catches of more than 0.73, 0.84, or 1.07% marlin for 1°, 2°, or 5° cells, and there would have a concomitant reduction in target species catch of 59.4, 63.1, or 66.5% respectively. Obviously, the temporal and spatial distribu- tion of the cells that would be eliminated with the current method for any particular thresh- old selected is of considerable interest. Figure 3 presents the distribution of cells by area and month that correspond to a threshold selection that would have eliminated 50% of the billfish bycatch from the o a T3 m i - CTj) fo-1 A-1' a^O a = 0 (1) The CTj and a^ are standard deviations for the mini- mum and maximum ages, respectively. The param- eter a determines the nonlinearity of the function, where o( b ) becomes linear in 6 as a tends to 0. Given the parameter vector 0=(a], o^, a) and observed age classes a, the age-error matrix q(a |6,0) is de- fined by g(a|i>,): JO) (2) where -^^^(0) is the discrete normal density function such that -m '■ab (*) = (3) In Richards et al. ( 1992), the assumed "true age" for a fish aged by multiple readers is the modal age among multiple readers. We used the mean age rounded to the nearest integer because the mode is not defined for two readings when the ages differ. For the reader and known-age data, the true age is the known age. A value for the maximum age A is required. For the reader and known-age data set, we set A equal to the maximum known age. For the pri- mary reader and tester data set, we set A equal to the maximum assigned age. The model of Richards et al. (1992) does not in- clude estimation of bias. The use of known ages al- lows bias to be estimated and incorporated in the ageing-error matrix. By including three additional parameters, Ericksen (1997) generalized the meth- ods of Richards et al. ( 1992) to include estimation of bias for tag recapture and known-age data. As with Equation 1, for a given true age 6, the bias jiib) of the observed age is defined by three parameters, j3j, /3^, and A, such that = (ct,, ct^, a, /3j, (i^. A) and Pih) = - ^_ -hb-h P^+(Pa-Pi\ _,,^_1, 1-e /^i+(/?.A-/?,)|3^; Equation 3 is modified to obtain ; X^O A = 0 (4) Heifetz et al.: Age validation and analysis of ageing error for Anoplopoma fimbria 259 •■ab (*) ^r£-ii*£iMii' ■2 ai6. ^2no{b) (5) To estimate the classification matrix defined by Equa- tion 2, maximum likelihood was used to estimate the parameters of the model. The likelihood (L) of the observed ages A given the true ages B is / .] L(A\B) = WY\q(a^j\b„e<-« B^oa n ixi — ^- -) — a — 1 — i — 1 — 1 — 1 — 1 — 1 — 1 — ^ XI X i 1 — 1 1 i i i- Table 1 Alternative cases for analysis of ageing error based on pri- mary reader and tester data (data set ll and reader and known-age data (data set 2). Case Data set Parameters Constraints 1 1 tr,, C7^. a — 2 1 CT,, CT, a = 0 3 4 5 6 2 2 2 2 CT,,fT^, a,/J,,/J,^,A CTi, CT,. a, /?,, P^ c,.a^.PyP,,.>^ o,.o,.p,.p. A = 0 a=0 A = 0; a = 0 sidered unreadable. Of the testers's misaged fish, most (70.1%) were misaged by 1 year: 10 of the 13 underaged fish and 9 of the 14 overaged fish. The largest discrepancy was 4 years; known age was 9 years and tester age was 13 years. Agreement between primary reader and tester ages was much greater than between the known and pri- mary-reader ages or the known and tester ages (Fig. 1). Primary reader and tester matched ages in 24 of the 44 (54.5%) specimens compared, indicating that they interpreted annuli similarly. Most discrepan- cies {n-12\ 60.0%) were discrepancies of one year. The primary reader tended to age the fish younger than the tester: 12 less than tester ages and 8 greater Only once did their ages differ by more than two years (known age=9; primary reader=9; tester=13). 260 Fishery Bulletin 97(2), 1999 Table 2 Comp arisen of pa rameter estimat es foi alternative models of ageing error for sablefis h based on primary read er and tester data (data set 1) an d reader and known-age data (data set 2). N urr ber of observations ( n i , maximum age (A), num ber of parameters (A^), minimum va ues of the inference function (/). and Akaike information criterion (AIC) are also given. Data Case set n A N f^i ('a « ft Pa A / AIC 1 1 88 13 3 0.399 2.464 -0.358 — — — 139.3 145.3 2 1 88 13 2 0.270 1.094 set to 0 — — — 143.1 147.1 3 2 92 9 6 0.253 1.792 0.341 -0.284 0.289 0.116 308.8 320.8 4 2 92 9 5 0.213 1.796 0.346 -0.191 0.307 set to 0 308.9 318.9 5 2 92 9 5 0.911 1.948 set to 0 -1.731 0.276 0.604 310.0 320.0 6 2 92 9 4 0.699 2.089 set to 0 ^0.340 0.433 set to 0 310.4 318.4 This study generally confirmed the criteria used for ageing young sablefish. Even though a high pro- portion offish were misaged, most fish were misaged by only one year. In addition, after reexamining the otoliths after the known ages had been revealed, the primary reader could reconcile the differences be- tween the known and assigned ages in most cases (25 of 31). Anderl and Heifetz- have described three types of misinterpretation of otolith patterns that resulted in most misages: 1) misinterpretation of an ambiguous check (i.e. false annulus) immediately following the first annulus; 2) misinterpretation in assessing whether the most recent years annulus had been formed; and 3) misinterpretation of mul- tiple checks on parts of the otolith (a check is a mark, growth zone, or part of a growth zone on an otolith that does not form annually but reflects various en- vironmental or physiological changes [Chilton and Beamish, 1982]). Analysis of ageing errors Table 2 summarizes the parameter estimates for various model specifications from the data in Figure 2. The estimates based on between-reader variabil- ity include only data where both primary reader and tester readings were available. The data consisted of/ = 44 fish and J = 2 readings per fish resulting in ;i := 88 observations. The estimates based on reader and known ages were derived from data where at least one reader determined an age. The data con- sisted of / = 44 fish with J = 2 readings per fish and 7=4 fish with J = 1 reading per fish resulting in n = 92 observations. For the estimates that included only between- reader variability, a model with all three parameters 14 1 13 1 12 ~ 11 1 ..- i 10 1 ..-■■■ a 9- 1 1 ..-••■ 1 1 S 8 1 ..2-' 1 " 7 1 . 2-' S 6 1 ..3--' 3 CD .4-' H 4 1 .■%■' 2 3- 3 ..3-' 2 • 2- ■ 1 ■ 1 2 3 4 5 6 7 8 9 10 11 12 Primary reader age (yr) ~ 14 1 £: 13 a, 12 1 en 11 1 ™ 10 1 1 1 £ 9 .2 ■2 8 1 1 ..■!•• 3 (13 7 CD ' 2 ..V 1 - 6 ..•3' 2 ? 5 2 ...2' 3 fO 4 1 .3' 3 1 E 3 1 ..2 2 ^ ? 3' 3 1 2 3 4 5 6 7 8 9 10 11 12 Known age (yr) 14 T 13 1 _ 12 S- \l 1 — 10 1 OJ 9 1 2.1 CO 0 1 1 .l" 1 7 1 . -2' QJ 6 1 1 ..2' 2 1 f; 5 1 ..1 ■ 2 ^ 4 1 ...5" 3 2 3 1 .-4 " 1 2 V 1 1 2 3 4 5 6 7 8 9 10 11 12 Known age (yr) Figure 2 Observed relation for sablefish between primary reader age and tester age, known age and primary reader age, and known age and tester age. Points are repre,sented by the frequency distribution of observations. A dotted line of slope 1 is included for reference in each panel. ^ Anderl, D., and J. Heifetz. 1999. An evaluation of ageing crite- ria for sablefish ba.sed on known age specimens. In prep. estimated (case 1 ) provided the best fit with the few- est parameters (i.e. lowest AIC values ). The AIC value for a model with or constrained to 0 (case 2) was nearly Heifetz et al,: Age validation and analysis of ageing error for Anoplopoma fimbria 261 f r 0.7 2.3 -*- Case 1 SD 1 -»- Case 4 SD > / 0.5 -*- Case 6 SD y^ 1 rd deviation (yr) -•- Case 4 bias / / f -•- Case 6 bias /^^^"V^ f 0.3 CD ^ vy / m ■o 55 /V / ' 0.1 ^ 0.8 9^^ •-0.1 0.3 ^ /L^^-^-^^ -0.3 , 1 2 3 4 5 6 7 8 9 10 11 12 13 14 Age (yr) Figure 3 Estimated standard deviation (SD) and bias of obser\'ed age of sable- fish for each true age based on primary-reader and tester ages (case 1) and reader and known ages (case 4 and case 6). Case refers to model specifications listed in Table 1. 2 units above the AIC value for case 1. For the reader and known-age data set, a model with a and X con- strained to 0 (case 6) provided the best fit with the fewest parameters. The AIC value for a model with only A constrained to 0 (case 4) was only 0.5 units above the minimum. The closeness of the AIC val- ues for these two cases indicates that case 6 provides only a slightly better model fit than case 4. The differ- ence between case 4 and case 6 is best understood by comparing estimated standard deviation and bias at age (Fig. 3). Case 4 results in a nonlinear relationship between standard deviation and age, and the standard deviation approaches an asymptote near age 9; whereas for case 6, the standard deviation increases linearly with age. Bias is linear for both cases, and case 6 has higher absolute bias than case 4 for most ages. Estimates of ageing errors based on comparison of known ages to reader ages were considerably higher than estimates obtained from between-reader vari- ability (Fig. 4). For example, for age 2-9 fish, accord- ing to primary reader and tester ages, the estimates of the probability of assigning the true age was 0.95- 0.46. In contrast, according to the estimates from reader and known ages, the probability was only 0.71-0.19. This discrepancy was expected because agreement between readers was considerably greater than between reader and known ages. Thus, use of between-reader agreement to assess ageing error may lead to a false sense of the true error. In conclusion, use of ageing errors based on known- age samples may help improve stock assessment of sablefish. Future analysis of ageing errors for sable- fish may require consideration of the time of year otoliths are taken because, as Anderl and Heifetz- have shown, ageing error may depend on the season when an age sample is taken. Precaution should be taken in extending our results to fish older than age 9. Results should be compared between stock assessments that use parameter estimates for the ageing-error matrix based on case 4 and case 6. If a sample is obtained that includes older known-age fish, the ageing-error matrix can be estimated for older fish. We expect such a sample to be available in the future as the fish tagged as juve- niles by Rutecki and Varosi ( 1997a) continue to be re- covered. For many species other than sablefish, knowm- age specimens are not available. For such species, vari- ability between readers may be the only data available to assess ageing error Such data are valuable for evalu- ating the precision and consistency of ageing criteria applied by different readers. Estimates derived solely from between-reader variability should be viewed as minimum estimates of ageing error. Acknowledgments Critical reviews by Dave Clausen, Jerome Pella, Mike Sigler, Jeff Fujioka, and two anonymous referees 262 Fishery Bulletin 97(2), 1999 Presumed tme s age 2 3 4 5 6 7 8 9 10 11 12 13 Observed age Actual true age 2 3 4 5 6 7 8 9 10 11 12 13 Observed age Figure 4 Estimated probability of observing a given age for a particular true age of sablefish estimated from primary-reader and tester ages (A, case 1) and from reader and known ages (B, case 6). Case refers to model specifications listed in Table 1. helped improve this manuscript. We thank JuHe Lyons for serving as the tester for this study and our co-workers that tagged numerous juvenile sablefish throughout southeastern Alaska. Our appreciation is also extended to the many observers and scien- tists who provided otoliths of recaptured sablefish. We also thank James lanelli for helping implement estimation of the ageing-error matrix. Literature cited Akaike, H. 1974. A new look at the statistical identification model. Institute of Electrical and Electronic Engineers (IEEE) Trans. Auto. Control 19:716-72.3. [Referred to in Richards et al. (1992).] Beamish, R. J., and D. E. Chilton. 1982. Preliminary evaluation of a method to determine the age of sablefish tAnoplopnina fimbria). Can. J. Fish. Aquat. Sci. .39:277-287. Beamish, R. J., G. A. McFarlane, and D. E. Chilton. 1983. Use of oxytetracyclme and other methods to validate a met hod of age determination for sablefish. In Proceedings of the second international sablefish symposium, p. 9.5-116. Alaska Sea Grant Rep. 83-8. Univ. Alaska. Fairbanks, AK. Chilton, D. E., and R. J. Beamish. 1982. Age determination methods for fishes studied by the groundfish program at the Pacific Biological Station. Can. Spec. Publ. Fish. Aquat Sci. 60, 102 p. Ericksen, R. P. 1997. Estimation of aging accuracy and precision, growth, Heifetz et a\:. Age validation and analysis of ageing error for Anop/opoma fimbria 263 and sustained yield of coastal cutthroat trout in South- east Alaska. M.S. thesis. Univ. Alaska Fairbanks, Juneau, AK. 126 p. Foumier, D., and C. P. Archibald. 1982. A general theory for analyzing catch at age data. Can. J. Fish. Aquat. Sci. 39:1195-1207. Kastelle, C. R., D. K. Kimura, A. E. Nevissi, and D. R. Gunderson. 1994. Using Pb-210/Ra-226 disequilibria for sablefish, Anoplopoma fimbria, age validation. Fish. Bull. 92:292- 301. Kimura, D. K. 1990. Approaches to age-structured separable sequential population analysis. Can. J. Fish. Aquat. Sci. 47:2364- 2374. Kimura, D. K., and J. J. Lyons. 1991. Between-reader bias and variability in the age-de- termination process Fish. Bull. 89:53-60. Lai, H.-L. 1985. Evaluation and validation of age determination for sablefish. pollock. Pacific cod and yellow fin sole; optimum sampling design using age-length key; and implications of aging variability in pollock. Ph.D. diss., Univ. Washing- ton. Seattle. 426 p. Lai, H. L., and D. R. Gunderson. 1987. Effects of ageing errors on estimates of growth, mor- tality and yield per recruit for walleye pollock tTheragra chalcugramma). Fish. Res. 5:287-302. McFarlane, G. A., and R. J. Beamish. 1995. Validation of the otolith cross-section method of age determination for sablefish {Anoplopoma fimbria) using oxytetracycline. In D. H. Secor, J. M. Dean. S. E. Campana (eds. ), Recent developments in fish otolith research, p. 319- 329. Univ. South Carolina Press, Columbia, SC. Methot, R. D. 1990. Synthesis model: an adaptable framework for analy- sis of diverse stock assessment data. In L.-L. Low (ed.). Proceedings of the symposium on application of stock as- sessment techniques to gadids, p. 259-277. International North Pacific Fisheries Commission (INPFC) Bull. 50. Richards, L. J., J. T. Schnute, A. R. Kronlund, and R. J. Beamish. 1992. Statistical models for the analysis of ageing error Can. J. Fish. Aquat. Sci. 49:1801-1815. Rutecki, T. L., and E. R. Varosi. 1997a. Migrations of juvenile sablefish. Anoplopoma fim- bria, in southeast Alaska. In M. E. Wilkins and M. W. Saunders (eds.). Biology and management of sablefish, Anoplopoma fimbria, p. 123-130. U.S. Dep. Commer. NOAA Tech. Rep. NMFS 130. 1997b. Distribution, age, and growth of juvenile sablefish, Anoplopoma fimbria, in southeast Alaska. In M. E. Wilkins and M. W. Saunders (eds.). Biology and manage- ment of sablefish, Anoplopoma fimbria, p. 45-54. U.S. Dep. Commer. NOAA Tech. Rep. NMFS 130. Sigler, M. F., S. A. Lowe, and C. R. Kastelle. 1997. Area and depth differences in the age-length rela- tionship of sablefish. Anoplopoma fimbria, in the Gulf of Alaska. In M. E. Wilkins and M. W. Saunders (eds). Bi- ology and management of sablefish. Anoplopoma fimbria. p. 55-63. U.S. Dep. Commer. NOAA Tech. Rep. NMFS 130. Tyler, A. V., R. J. Beamish, and G. A. McFarlane. 1989. Implications of age determination errors to yield estimates. In R. J. Beamish and G. A. McFarlane (eds.). Effects of ocean variability on recruitment and an evalua- tion of parameters used in stock assessment models, p. 27- 35. Can. Spec. Publ. Fish. Aquat. Sci. 108. 264 Abstract.— A manned submersible was used in the eastern Gulf of Alaska in 1992 to obser\'e spatial distributions and habitats of shortraker rockfish, Sebastes borealis, and rougheye rock- fish, S. aleutianus, on the continental slope at 262-365 m depths. Observa- tions of these two species were com- bined because distinguishing between them was not always possible from the submersible. A seafloor area of 104,900 m^ was surveyed at 15 dive sites, and 646 shortraker and rougheye rockfish were observed. Densities were 0.0 to 14.8 rockfish/1000 m- (mean, 5.8 rock- fish/1000 m-l. Of the 646 rockfish, 115 were observed above bottom and 531 were on the bottom. The above-bottom rockfish were descending slowly to the seafloor and became sedentary when they contacted the seafloor. Approxi- mately two-thirds of the rockfish were in groups; 82 of the 113 groups con- tained 2 or 3 rockfish. and only 2 groups had more than 12 rockfish. Rockfish were associated w'ith 20 of the 22 sub- strates encountered. Soft substrates of sand or mud usually had the greatest densities of rockfish, whereas hard sub- strates of bedrock, cobble, or pebble usually had the least densities. Habi- tats containing steep slopes and numer- ous boulders had greater densities of rockfish than habitats with gradual slopes and few boulders; 52 rockfish lay against boulders. According to catch rates from bottom-trawl surveys, popu- lations of shortraker and rougheye rockfish may be underestimated be- cause of the above-bottom distribution of these rockfish and their use of steep- slope boulder habitats. Distribution and abundance of shortraker rockfish, Sebastes borealis, and rougheye rockfish, S. aleutianus, determined from a manned submersible Kenneth J. Krieger Auke Bay Laboratory, Alaska Fisheries Science Center National Marine Fisheries Service, NOAA 11305 Glacier Highway Juneau, Alaska 99801-8626 E mail address Ken Kneger g noaa gov Daniel H. Ito Alaska Fisheries Science Center National Marine Fisheries Service, NOAA 7600 Sand Point Way N E Seattle, Washington 98115-0070 Manuscript accepted 20 May 1998. Fishery Bulletin 97:264-272 (1999). Shortraker rockfish, Sebastes borea- lis, and rougheye rockfish, S. aleu- tianus, occur in commercial quanti- ties from northern Washington throughout Alaska (Allen and Smith, 1988). Both species are similar in appearance and share similar life- history patterns. These rockfish at- tain maximum total lengths of about 100 cm (Kramer and O'Con- nell, 1986) and have been aged at more than 120 years (Chilton and Beamish, 1982). Their bathynietric range is 25-875 m (Allen and Smith, 1988), and their lengths at 50^, ma- turity are 43.97 cm for rougheye rock- fish and 44.90 cm for shortraker rock- fish (McDermott, 1994). They appar- ently share similar habitats. During the 1996 Gulf of Alaska triennial sur- vey, 89*^ of the trawl hauls contain- ing shortraker rockfish also contained rougheye rockfish. Shortraker and rougheye rockfish are harvested with bottom trawls and longlines in the Gulf of Alaska. Until the mid 1980s, they were mainly bycatch species caught dur- ing longlining for halibut, Hippo- glossus steiwlepis, and sablefish, Anoplopoma fimbria, and during bottom trawling for more abundant rockfish species. Before 1991, shortraker and rougheye rockfish were combined with 18 other rock- fish species and managed as "slope rockfish." Since 1991, shortraker and rougheye rockfish have been managed as a separate subgroup because fishermen target these highly valued species. For example, shortraker and rougheye rockfish made up 337f of the commercial rockfish catch in the eastern Gulf of Alaska in 1990 but made up only 14*7^ of the estimated rockfish bio- mass (Heifetz and Clausen, 1991). Catch quotas of shortraker and rougheye rockfish are based prima- rily on population estimates derived from catch rates of bottom-trawl surveys ( Heifetz etal., 1996). These estimates are suspect because the catch efficiency of bottom trawls on these species is unknown and only certain types of habitats can be sampled with bottom trawls. Catch rates from bottom-trawl sur- veys are converted to biomass esti- mates by assuming a 100'^ sampling efficiency for the area swept by the trawl. The area that is swept is de- termined as the distance between the wingtips of the net and the distance Krieger and Ito: Distribution and abundance of Sebastes borealis and 5. aleutianus 265 the net is towed. The samphng efficiency could be less than 100% if fish are distrib- uted above the headrope, protected by structures such as boulders, swim out of the path of the net, or escape under the net. Conversely, if fish are herded into the trawl by the bridles and doors and do not escape above or under the net, the sam- pling efficiency may be greater than 100%. Typically, smooth (trawlable) substrates are sampled during trawl surveys. Catch rates fi-om trawlable substrates are then applied to all substrates for estimating bio- mass. Bottorn-trawl surveys may not pro- vide reliable biomass estimates of short- raker and rougheye rockfish because 1 ) the sampling efficiency may not be 100% for the distance between the wingtips of the net, 2) these species may use untrawlable substrates at a different rate than they do trawlable substrates, and 3) the sampling fi-equency may not be sufficient, depend- ing on the distribution patterns of the tar- get species. Minimal information is available on the distribution of shortraker and rougheye rockfish. Fishermen report that rockfish school above bottom in steep-slope areas. From a manned sub- mersible Krieger (1992) observed 20 shortraker rockfish on the continental shelf; these fish were in contact with the seafloor and were distributed as solitary individuals on shallow-sloped, smooth habitat. Shortraker rock- fish were observed only at sites where boulders were common, and six of the fish were found next to boul- ders 0.5-1.5 m in diameter (Krieger, 1992). Catches of shortraker and rougheye rockfish during longline surveys indicate they are most abundant on the up- per continental slope at 300-400 m depths (Sigler and Zenger, 1994), but most of this substrate is con- sidered untrawlable and is seldom sampled during bottom-trawl surveys. For example, only eight trawl hauls were completed along the 500-km continental slope in southeastern Alaska during the last four bottom-trawl surveys ( 1987, 1990. 1993, 1996). We need to understand the distribution and habi- tats of shortraker and rougheye rockfish to assess them effectively with bottom trawls or other sam- pling gear. In this study, a manned submersible was used to observe their spatial distributions and habi- tat associations. These species were also quantified from the submersible for comparison with abundance estimates from bottom-trawl surveys. The two spe- cies were combined and are referred to as rockfish in 137 W w 129 W ^ Alaslce Study Area ..^^ 1 "% Gulf of Alaska Figure 1 Submersible survey sites for shortraker and rougheye rockfish in the eastern Gulf of Alaska in May 1992. this paper because distinguishing between them was not always possible from the submersible. Materials and methods Study area This study was conducted in May 1992 on the upper continental slope in the eastern Gulf of Alaska between lat. 56=10' and 58°10'N (Fig. 1). This region has consis- tently produced high catch rates of shortraker and rougheye rockfish during the annual longline surveys in the Gulf of Alaska (Rutecki et al., 1997). The study area spanned more than 200 km to include a variety of habitats. Distances separating adjacent sites ranged from 0.2 to 84.2 km. Dives were conducted during day- light, between 0600 and 1900 hours. Submersible The two-man submersible Delta was chartered for all dives. This battery-powered submersible is 4.7 m 266 Fishery Bulletin 97(2), 1999 long, dives to 365 m, and travels 2-6 km/h for 2-4 h. It is equipped with halogen lights, internal and ex- ternal video cameras, magnetic compass, directional gyro compass, underwater telephone, and transpon- der that allow tracking of the submersible from a sur- face vessel. On each dive, the submersible descended to 265-365 m and then usually traveled parallel to the shelf break, followed by up-slope travel to less than 300 m before it ascended. The surface vessel recorded LORAN fixes at the beginning and end of a dive, and every 1-5 min during a dive. The submersible traveled 1.0-2.5 km/h, depending on the slope and ruggedness of the terrain and the magnitude and direction of the current. The degree of slope determined how observa- tions were made from the submersible. When the slope was less than 60°, the submersible remained in con- tact with the seafloor while the scientist viewed the water column parallel to the seafioor through a star- board porthole 0.5 m above the base of the submers- ible. When the slope was greater than 60°, the sub- mersible traveled 2-3 m away from the seafloor while the scientist viewed the water column almost per- pendicular to the seafloor through portholes on the starboard side and bow. The pilot sat above the ob- server in a tower with a panoramic view and assisted in locating fish, especially above the submersible. The submersible lights provided constant illumination. Data analysis Observations of rockfish and their habitat were audio- tape- and videotape-recorded for subsequent analysis and verification. The senior author reviewed all video tapes by 1-min segments ( 16-42 m travel distances ), and four habitat parameters were estimated: 1 ) main substrate, 2) secondary substrate, 3) slope, and 4) boulder abundance. The main substrate made up 50- 100% of the substrate, whereas the secondary sub- strate made up 10-50'^ of the substrate. Substrates consisted of mud, sand, pebble, cobble, and bedrock. Granular size used to separate pebble and cobble was 2.5 inches (64 mm), and to separate cobble and boul- ders was 10 inches (256 mm). Mud would stay sus- pended when disturbed by the submersible, whereas sand would not. Size references for classifying pebble, cobble, and boulders included the known length and width of a video frame as well as the known, uni- form size of invertebrates such as sea stars and shrimp. The slope was classified into four categories: 1 = 0-5°, 2 = 6-20 \ 3 = 21-45°, and 4 = 46-90 . Slope classification was based on estimates by the pilot and on the view from a downward-aimed, mounted video camera. Boulder abundance was classified into five categories: 0 - absent, 1 = scarce, 2 - scattered patches, 3 = common (usually in view), and 4 = abun- dant (always in view). For each site, we calculated the average slope, average boulder abundance, and the densities of rockfish associated with each sub- strate. For all sites combined, we calculated rock- fish densities associated with each substrate, and the percentage of rockfish associated with each slope category and each boulder category. Rockfish observations included number, size, grouping behavior, above-bottom distribution, and movements. Three sizes of rockfish were estimated visually: small (<30 cm), medium (30-60 cm), and large (>60 cm). The observers had used laser beams to measure rockfish lengths from submersibles and were, therefore, experienced in sizing them. Rock- fish were considered grouped if they were within 5 m of each other. Densities were estimated from counts of rockfish and the total seafloor area surveyed. The surveyed area was the distance the submersible trav- eled (0.3-1.8 kin/dive) multiplied by the estimated viewing distance from the submersible (4-10 m, de- pending on water clarity). These estimates were cali- brated against sonar readouts when the seafloor be- came visible during descents and when the seafloor disappeared from view during ascents. Estimated distances were within 1 m of true distances accord- ing to sonar readouts and distance calibrations in previous studies (Krieger, 1993). Rockfish movement rates were based on the estimated distance moved during a specific time period. Results Submersible dives Fifteen submersible dives were completed and 104.900 m- of seafloor was surveyed at 262-365 m depths (Table 1). Rockfish densities ranged from 1.2 to 14.8 rockfisli/1000 m" (mean, 5.8/1000 m-) at the 14 sites where they were observed. Of the 646 rockfish observed, 188 were small, 289 medium, and 169 large. Above-bottom and on-the-bottom behavior We observed 115 rockfish 1-10 m above bottom and 531 on the bottom (Table 2). Sites 5 (80 above-bot- tom rockfish) and 13 (19 above-bottom rockfish) ac- counted for 86% of the above-bottom rockfish. Above- bottom rockfish were medium-size (108 rockfish) or large (7 rockfish), and they were observed descend- ing at less than 10 m/min without detectable move- ments of fins or body. They would contact the seafloor without disturbing sediments and would orient broad- side to the current, which would tilt them 10-45° (Fig. 2). Currents were less than 1.0 km/h at all sites. Krieger and Ito: Distribution and abundance of Sebastes borealis and 5. aleutianus 267 Table 1 Number and density of shortraker and rougheye rockfish <30 cm; Medium = 30-60 cm; Large = >60 cm. at 15 submersible dive sites in the eastern Gulf of Alaska, 1992. Small = Site Depth (m) Surveyed area (1000 m2) Number of rockfish Density of rockfish (no./lOOO m-) Total Small Medium Large 1 365-262 8.1 25 16 7 2 3.1 2 323-270 1.8 0 0 0 0 0.0 3 346-285 3.5 21 10 6 5 6.0 4 362-262 10.1 60 30 26 4 5.9 5 365-292 13.0 192 24 165 3 14.8 6 365-270 7.9 43 16 19 8 5.4 7 365-270 4.0 5 3 0 2 1.2 8 365-275 7.5 45 7 9 29 6.0 9 365-300 13.5 16 0 3 13 1.2 10 365-270 1.7 21 10 6 5 12.4 11 365-304 7.3 64 28 15 21 8.8 12 365-280 2.0 17 3 3 11 8.5 13 365-320 16.3 98 29 23 46 6.0 14 365-288 5.8 34 9 5 20 5.9 15 365-285 2.4 5 3 2 0 2.1 Totals 104.9 646 188 289 169 Table 2 The number of shortrake - and rougheye rockfi sh observed on the bottom, above bottom, solitary, and grouped, and their distribu- tion by group size at 15 submersible dive sites in the eastern Gulf of Alaska, 1992. Fish Fish Soli- No. of Group size on above tary fish No. of Site bottom bottom fish in groups groups 2 3 4 5 6 7 8 9 10 11 12 — 15 32 1 24 1 15 10 4 2 2 2 0 0 0 0 0 3 21 0 14 7 2 1 1 4 59 1 30 30 11 6 3 1 1 5 112 80 35 157 27 9 6 2 3 1 112 11 6 42 1 19 24 9 6 2 1 7 5 0 5 0 0 8 39 6 10 35 10 4 4 1 1 9 15 1 9 7 3 2 1 10 21 0 6 15 6 3 3 11 62 2 37 27 11 8 1 2 12 17 0 7 10 3 1 2 13 79 19 21 77 22 13 1 3 1 1 2 1 14 30 4 22 12 5 4 1 15 5 0 5 0 0 Totals 531 115 235 411 113 59 23 11 6 4 0 2 2 112 11 Of the 531 rockfish on the bottom, only 21 moved when passed or approached by the submersible; 5 moved 0.5-1.0 m above the seafloor and swam less than 5 m away at a speed of less than 1 km/h, and 16 moved less than 2 m along the seafloor by drifting or by slight movements of fins. The submersible drifted against several large rockfish, which did not respond as they were pushed along the seafloor. 268 Fishery Bulletin 97(2), 1999 n-ixi} 'ii^^mm Figure 2 Shortraker or rougheye rockfish tilted with the current while lying on the seafloor in the eastern Gulf of Alaska in May 1992. Observed from a submersible. Spatial distribution Of the 115 above-bottom rockfish. 106 were grouped with at least one other rockfish as they descended together to the seafloor. They contacted the seafloor within 5 m of at least one other rockfish in the group. Because these fish maintained a <5-m spacing from each other, all rockfish within 5 m of other rockfish on the seafloor were considered grouped. Twelve sites had 411 grouped fish, and 14 sites had 235 solitary fish (Table 2). Most groups were small; 82 of the 113 groups contained only 2 or 3 rockfish, whereas only two groups contained more than 12 rockfish. Twenty- six pairs of rockfish consisted of a small and medium- size fish separated by less than 0.5 m. Substrate associations Twenty-two combinations of primary and secondary substrates were encountered at the 15 sites (Table 3). Substrates changed frequently within each site, averaging 6.9 substrates/site. Rockfish were associ- ated with 20 of the substrates, but no consistent pat- tern of association was observed within a site. F'or example, the 8 greatest densities (>16 rockfish/1000 m-) included 7 different substrates, the 23 lowest densities (no rockfish) included 13 different sub- strates, and 3 substrates had both more than 16 rock- fish/1000 m^ and no rockfish. For all sites combined, the most abundant sub- strate was cobble with sand (17.3%), whereas 7 of the 22 substrates made up 1.0% or less each of the total substrate (Table 4). The greatest densities of rockfish were usually associated with soft substrates. Sand with mud had the greatest average density (9.1 rockfish/lOOO m-), and 7 of the 10 greatest densities were associated with primary substrates of sand or mud. Nine of the 12 lowest densities of rockfish were associated with primary substrates of cobble, rock, or pebble. Boulder and slope associations Average boulder indices ranged from 0.0 to 2.1 (Table 3 ) and were not highly correlated (r=0.30 ) with rock- fish densities. For all sites combined, high-abundance boulder habitat contained greater densities of rock- fish than did low-abundance boulder habitat (Fig. 3). Habitats where boulders were absent (index=l) or rare (index=2) were encountered 67% ofthe time and had 56% of the rockfish, whereas habitats where boulders were more abundant (index=3, 4, or 5) were encountered 33% ofthe time and had 44% ofthe rock- fish; 52 rockfish were found lying against boulders ( F'ig. 4). The only site without boulders (site 2) was the only site not containing rockfish. Average slope indices ranged from 0.9 to 3.7 and were not highly correlated (r=0.56) with rockfish Krieger and Ito: Distribution and abundance of Sebastes borealis and S. aleutlanus 269 Table 3 Densities of s hortraker and rougheye rockfish associated w ith s pecific substrates at 15 subme rsib e dive site s in southeastern Alaska, 1992 and in dexes of boulder ab undance and topographj (slope). Indexes are averages of 1 -min video segments, where boulders were absent (01, rare (1), s cattered patches (2 , common (3), or abundant (4l, and slope s were 0- 5(1), 5-20= (2), 20-45° (3), or 45-90° (4). M = mud; S = sand; C = cobble; P = pebble ;R = rock M-S = mud and sand; M-C = mud and cobble, etc. Boul- der Slope Rockfish densities ( no./lOOO m^) and associated substrates Site index index M M-S M-C M-P M-R S S-M S-C S-P S-R P-S P-M P-C C C-S C-M C-P C-R R R-S R-M RyP 1 1.1 2.0 3.1 3.1 2.1 4.1 3.1 3.1 2.1 2.2 2 0.0 2.0 0.0 0.0 0,0 0.0 0.0 3 0,7 3.0 4.0 12.8 12.0 8.0 0,0 1.0 0,0 4 1.1 3.0 0,0 11.8 12.8 0.8 3.0 5.9 8.9 7.9 9.6 4,9 5 1.7 2.9 5.1 22.3 70.9 9.7 16,9 6 1.2 . 3.5 0.0 8.7 4.0 1,9 13,0 5.8 5.8 0.0 0.0 17.3 6.7 2.9 7 1.6 2.0 0.8 3.2 1.1 0,8 8 2.1 2.0 3.0 0.0 6.5 6.0 6.0 7.4 0.0 9 0.8 0.9 1.0 1.3 3.9 1.6 0.0 0.5 1.1 0.0 0.0 0,0 10 0.4 3.7 13.7 25,1 0.0 11 0.7 3.7 6.5 25.8 12.9 0.0 6.5 9.0 3.2 17.7 6.5 12 1.1 3.0 11.7 7.6 13 1.7 3.2 14,0 3.0 5.4 6.0 5.5 9.5 7.5 3.0 3.0 3.0 14 1.9 2.9 11.7 28.0 1.7 0.0 4,9 5.9 15 1.2 2.4 0.0 4.5 0.0 4,5 0.0 4.5 densities (Table 31, For all sites combined, habitats with slopes greater than 20° contained greater den- sities of rockfish than those with slopes less than 20° (Fig. 5). Slopes less than 20° were encountered 37% of the time and had 24% of the rockfish, whereas slopes greater than 20° were encountered 63% of the time and had 767f of the rockfish. Submersible counts versus trawl catch rates During the last four bottom-trawl surveys (1987, 1990, 1993, 1996), eight trawl hauls were completed on the continental slope in southeastern Alaska and 234,100 m" of seafloor was sampled. The mean catch rate was 3.2 rockfish/1000 m" for the 8 trawl hauls, compared with the mean observation rate of 5.8 rock- fish/1000 m- at the 15 dive sites. Catch rates exceeded 5.0 rockfish/1000 m- at only 2 of the 8 trawl sites, whereas observation rates exceeded 5.0 rockfish/1000 m^ at 10 of the 15 dive sites. Discussion Because the two rockfish species were combined in this study, we did not determine whether shortraker rockfish behavior differs from rougheye rockfish be- havior. They appear to share the same habitats, -based on bottom-trawl sampling and observations from the submersible, but differences may exist in Table 4 The percentage of each seafloor substrate and density of shortraker and rougheye rockfish associated with each sub- strate observed during 15 submersible dives n southeastern Alaska, 1992. Substrate Substrate Rockfish type (%) (no./lOOO m-l sand-mud 1.0 19.1 cobble-sand 17,3 9.7 mud-cobble 1,7 9.3 sand 6,5 8.4 mud-pebble 5,0 7.0 sand-cobble 10.0 6.7 rock-mud 2.0 6.7 mud-sand 7.5 6.1 rock-sand 3.6 6.0 sand-pebble 10.5 5.5 pebble-mud 2.4 5.5 pebble-sand 11.5 3.7 cobble-pebble 7.7 3,5 rock 0.6 3.0 rock-pebble 1.0 3.0 pebble-cobble 6.0 2.5 cobble-rock 0.5 2.1 sand-rock 0.6 1.5 mud 2.1 0.9 mud-rock 1.1 0.8 cobble-mud 0.8 0.0 cobble 0.4 0.0 270 Fishery Bulletin 97(2), 1999 None Occasional Scattered Common Abundant Boulder abundance Figure 3 Fwe levels of boulder abundance observed on the seafloor, and the percent occurrence of habitat and percent occurrence of shortraker and rougheye rockfish associated with each level. Observations are from a submersible in the eastern Gulf of Alaska, 1992. their above-bottom distributions, grouping behavior, and use of boulders. The 150 rockfish observed above bottom were at nine different dive sites and all were descending, indicating they were reacting to the submersible. They were probably seeking the seafloor in response to the submersible. A diving response to trawl and vessel disturbances has been noted for other species of offshore rockfish (Kieser et al., 1992). The protec- tion provided by the seafloor may explain their re- luctance to move when they are on the seafloor and approached by the submersible. Assuming rockfish were descending in response to the submersible, the proportion of rockfish observed above bottom is prob- ably a minimum estimate because some had prob- ably reached the seafloor before they were viewed from the submersible. The 80 rockfish above bottom at site 5 indicate that a high percentage of rockfish move above the seafloor, although the frequency and duration of their movements are unknown. Rockfish may move above bottom to capture prey such as the squid and lantern fish (Myctophidae) on which they are known to feed (Yang, 1993, 1996). About two- thirds of the rockfish were in groups of 2-6 fish; only two groups contained more than 12 fish. The reason for the close pairing of a small and medium rockfish is unknown; it is probably not related to mating be- cause female shortraker and rougheye rockfish shorter than 30 cm are not mature (McDermott, 1994). The spatial distribution of rockfish varied within dive sites. This variability can be partially explained by their grouping behavior and by their habitat as- sociations. Rockfish were associated with most of the habitats encountered, but the greatest densities were associated with soft substrates, frequent boulders, and slopes greater than 20°. Their association with soft substrates may be prey related. Pandalid shrimp and hippolytid shrimp, which concentrate on soft substrates, were the main prey of rougheye rockfish examined from the Gulf of Alaska (Yang, 1993, 1996) and from the Aleutian Islands (Yang, 1996). The as- sociation of rockfish with boulders in this study and on the continental shelf ( Krieger, 1992 ) indicates that boulders are important for these species. Perhaps boulders are a necessary habitat feature, because shortraker and rougheye rockfish were absent at the one site without boulders in this study and at the three sites without boulders in a previous study (Krieger, 1992). These species may use boulders as territorial markers, to avoid currents, or to capture prey. Rockfish were least abundant on shallow-slope habitat (<5°) in this study, and Krieger (1992) ob- served shortraker rockfish at three sites where the slope was 3-12 " but none at six sites with slopes less than 2°. Steep slopes may provide relief fi-om currents. The mean observation rate from the submersible was about twice the mean catch rate from bottom- trawl surveys, probably because of the limited habi- tats sampled during trawl surveys. Bottom trawling may be effective for sampling shortraker and rougheye rockfish on low-relief habitats because these rockfish descended and remained on the sea- Krieger and Ito: Distribution and abundance of Sebastes borealis and 5. a/eutianus 271 Slope (degrees) Figure 5 Four levels of slope observed on the seafloor, and the percent occurrence of habi- tat and percent occurrence of shortraker and rougheye rockfish associated with each level. Observations are from a submersible in the eastern Gulf of Alaska, 1992. floor in response to the submersible. However, some above-bottom rockfish may not be captured because of their slow rates of descent. Bottom-trawl assess- ment gear is not designed to sample the steep-slope, boulder habitats occupied by these species, and the few trawl hauls that are completed in boulder habi- tat likely do not sample rockfish associated with boul- ders. An assessment method is needed that addresses the wide range of habitats, above-bottom distribu- tion, and grouping behavior of shortraker and 272 Fishery Bulletin 97(2), 1999 rougheye rockfish. Longline gear and bottom trawls with large rollers can sample rugged habitat, and studies are currently underway to determine the ef- ficiency of these gears for sampling shortraker and rougheye rockfish. Acknowledgments We thank the crews of the submersible Delta and the support vessel Cavalier for completing safe and successful dives. We also thank John Karinen for his assistance aboard the Cavalier, and Scott Johnson and Jon Heifetz for their advice and assistance in writing this paper. Literature cited Allen, J. M., and B. G. Smith. 1988. Atlas and zoogeography of common fishes in the Bering Sea and northeastern Pacific. U.S. Dep. Commer.. NOAA Tech. Rep. NMFS 66, 151 p, Chilton, D. E., and R. J. Beamish. 1982. Age determination methods for fishes studied by the groundfish program at the Pacific Biological Station. Can, Spec. Publ. Fish. Aquat. Sci. 60, 102 p. Heifetz, J., and D. M. Clausen. 1991. Slope rockfish. Chapter 5 in Stock assessment and fishery evaluation report for the 1992 Gulf of Alaska ground- fish fishery, p. 140-161. North Pacific Fishery Management Council, PO. Box 103136, Anchorage, AK 99.510. Heifetz, J., J. N. lanelli, and D. M. Clausen. 1996. Slope rockfish. In Stock assessment and fishery evaluation report for the 1997 Gulf of Alaska groundfish fisher.v, p. 230-270. North Pacific Fishery Management Council. PO. Box 103136. Anchorage. AK 99510. Kieser, R., B. M. Leaman, P .K . Withler, and R. D. Stanley. 1992. W. E. Ricker and Eastward Ho cruise to study the effect of trawling on rockfish behaviour. October 15-27. 1990. Can. Manuscr. Rep. Fish. Aquat. Sci. 2161, 84 p. Kramer, D. E., and V. M. O'Connell. 1986. Guide to northeast Pacific rockfishes, genera Sebastes and Sebastolobus. Alaska Sea Grant College Program, Univ. Alaska. Fairbanks. AK. Marine Advisory Bull. 25. 78 p. Krieger, K. J. 1992. Shortraker rockfish. Sebastes borealis, observed from a manned submersible. Mar Fish. Rev. 54(4):34-37. 1993. Distribution and abundance of rockfish determined from a submersible and by bottom trawling. Fish. Bull. 91:87-96. McDermott, S. F. 1994. Reproductive biology of rougheye and shortraker rockfish. Sebastes aleutianus and Sebastes borealis. M.S. thesis, Univ. Washington, Seattle, WA, 76 p. Rutecki, T. L., M. F. Sigler, and H. H. Zenger Jr. 1997. Data report: National Marine Fisheries Service longline surveys. 1991-96. U.S. Dep. Commer.. NO.AA Tech. Memo. NMFS-AFSC-83, 64 p. Sigler, M. F., and H. H. Zenger Jr. 1994. Relative abundance of Gulf of Alaska sablefish and other groundfish based on the domestic longline survey, 1989. U.S. Dep. Commer., NOAA Tech. Memo. NMFS- AFSC-40. 79 p. Yang, M-S. 1993. Food habits of the commercially important ground- fishes in the Gulf of Alaska in 1990. U.S. Dep. Commer. NOAA Tech. Memo. NMFS-AFSC-22, 150 p. 1996. Diets of the important groundfishes in the Aleutian Islands in summer 1991. U.S. Dep. Commer, NOAA Tech. Memo. NMFS-AFSC-60. 105 p. 273 Abstract.— Concentrations of lipids and protein were measured in embryos during gestation in two species of vi- viparous rockfishes off the central Cali- fornia coast. Total lipids and protein de- clined linearly through embryonic maturation in semipelagic yellowtail rockfish, Sebastes flavidus, and pelagic shortbelly rockfish, Sebastes jordani. Energetically, lipids were the predomi- nant source of energy for embryonic development in both species, but lipid and protein catabolism was signifi- cantly greater for yellowtail rockfish. Total lipids, protein, and lipid class composition were measured during embryonic maturation in three popu- lations of shortbelly rockfish, located at Ascension, Pioneer, and Bodega subma- rine canyons, to determine intraspecific variability of nutritional dynamics. Triacylglycerols and polar lipids (mostly phospholipids), the predominant lipid classes in all maturation stages, were depleted through embryonic develop- ment, Steryl or wax esters and choles- terol also declined, but were in much lower concentrations. The goodness-of- fit of linear regressions for protein, to- tal lipid, and lipid classes by stage of embryonic maturation allowed estima- tions of their concentrations at birth, thus providing a measure of nutritional condition, or qualitative reproductive success. Analyses determined that there were significant differences in metabolism and estimated concentra- tions at birth of nutrients between the two species and among the shortbelly rockfish populations, indicating differ- ential potential for survival during early planktonic life stages until favor- able feeding conditions occur Results suggest that the contribution of indi- vidual populations to the diversity of metapopulations or year classes may be influenced by the nutritional condition of larvae at birth. Nutritional dynamics during embryonic development in the viviparous genus Sebastes and their application to the assessment of reproductive success R. Bruce MacFarlane Elizabeth C. Norton Tiburon Laboratory, Southwest Fisheries Science Center National Marine Fisheries Service, NOAA 3150 Paradise Drive Tiburon, California 94920 E-mail address (for R, B. MacFarlane) Bruce MacFarlane a noaa gov ^Manuscript accepted 25 June 1998. Fish. Bull. 97:273-281 1 1999). Variability of annual recruitment to marine fish stocks has been attrib- uted to several factors, including the quantity and quality of progeny produced and the influences of en- vironmental conditions on subse- quent survival. Although either fac- tor could be effective alone, at times they may interact. Poor reproduc- tive success, manifested by low fe- cundity or unhealthy progeny, or both, occurring at times of unfavor- able environmental conditions, may lead to particularly weak year classes. There is evidence that fac- tors operating at or near the time of spawning aifect year-class strength in rockfishes (genus Sebastes) along the west coast of North America. Year-class failures of two species of rockfish were found at the extremes of the physical environmental spec- trum off the central coast of Cali- fornia during years of very low or very high upwelling (Ralston and Howard, 1995). The data from that investigation suggested that year- class strength of yellowtail rockfish {Sebastes flavidus) and blue rock- fish (Sebastes mystinus) was related to environmental conditions and was established before the pelagic juvenile stage. Further, annual variation in fecundity does not ap- pear to be great enough to account for the variation seen in fishery re- cruitment of yellowtail rockfish ( Eldridge and Jarvis, 1995 ) or other species (Shepherd and Gushing, 1980). Thus, maternal processes contributing to the health or fitness of embryos or early larvae, in addi- tion to the more often assessed en- vironmental factors, may signifi- cantly influence survival and year- class strength. The assessment of reproductive success includes the determination of both the quantity and quality of progeny because year-class strength may be influenced by the health of newborns as well as by the number produced. Although fecundity is of- ten used as a measure of reproduc- tive success and has been related to biological and environmental fac- tors (Blaxter, 1969), less attention has been given to the assessment of the qualitative aspects of repro- ductive success. This is due, in part, to the difficulty of determining which variables or processes are valid measures of egg, embryo, or larval health. Various measures have been proposed, such as egg size (Blaxter and Hempel, 1963), histo- logical criteria (Theilacker, 1978), and biochemical analyses, including nucleic acids (Buckley, 1984), en- zyme activity (Clarke et al., 1992), and biochemical composition (re- viewed by Ferron and Leggett, 1994). All have merits; however, fac- tors such as the degree of relation- 274 Fishery Bulletin 97(2), 1999 ship to growth or survival potential, ease or cost of analysis (or both), operational feasibility, or the abil- ity to assess adequate numbers of replicates for sta- tistical validity often diminish their utility in rou- tine assessments of reproductive success. Consider- ation of when, during egg and embryonic develop- ment, valid estimates can be made is also critical to accurate assessment of success. Ideally, evaluation at hatching or birth would provide the most accu- rate determination of health and survival potential, but this event is of very short duration, and obtain- ing enough individuals to characterize a population or species is difficult. The nutritional status, or energy content, of em- bryos and lai-vae has a clear relationship to growth and survival potential because the amount of meta- bolically available energy establishes the duration of survival in the environment until adequate food becomes available. Starvation during the early stages of development has been considered a major source of mortality or reduced fitness and may contribute to fluc- tuations in year-class strength (May, 1974; Shepherd and Gushing, 1980; Rissik and Suthers, 1996). Low and unstable abundance of prey typifies the environment off central California during winter and early spring (Ainley et al., 1993), the period when many species of viviparous rockfish give birth (Wyllie Echeverria, 1987). Pelagic larvae rely on endogenous nutritional reserves to supplement sparse forage until the biologically productive annual upwelling season begins, usually in late March or April (Bakun, 1975). This study describes protein and lipid metabolism during embryonic development in two species of the viviparous genus Sebastes. A procedure is presented to estimate the lipid and protein composition at par- turition, or birth, which provides a measure of nu- tritional status and thus may be related to the prob- ability for survival at the earliest free-living stage. Nutritional status at birth is presented for yellow- tail rockfish and three populations of shortbelly rock- fish (Sebastes Jordani), thereby providing compari- sons between congeners occupying different ecologi- cal niches and among populations separated spa- tially. These data are the first documentation of lipid class composition in eggs, embryos, and larvae of a viviparous marine teleost. Materials and methods Female S. flavidus and S. jordani were obtained fi-om January to March, the period of reproductive devel- opment spanning late vitellogenesis to parturition, at locations off the central California coast. Yellow- tail rockfish were captured by hook-and-line at Cordell Bank (38°01'N 123°25'W), a marine bank 37 km west of Pt. Reyes, at depths ranging from 50 to 150 m. Shortbelly rockfish were collected by trawl at 150 to 200 m depth in the proximity of three subma- rine canyons: Bodega Canyon (ca. 38°13'N 123°22'W), Pioneer Canyon (ca. 37'^24'N 122-52'W), and Ascen- sion Canyon (ca. 37°01'N 122°25'W). Fish were held on ice for up to 12 h until examination. Morpho- metries (e.g. standard length; body and liver weight) were recorded and ovaries excised, weighed, and stored at -70°C. The stage of oocyte or embryonic development was identified by microscopy according to the classification scheme of Yamada and Kusakari ( 1991) for Sebastes schlegeli (Table 1). Table 1 Embryo maturation stages (EMS) in Sebastes flavidus and Sebastes jordani . EMS Description 1 Late vitellogenic or migratory nucleus oocyte 2 Formation of germ disc 3 2-celled 4 4-celled .5 8-celled 6 16-celled 7 32-celled 8 64-celled 9 Morula 10 Early blastula 11 Late blastula 12 Beginning of epiboly 1.3 Early gastrula 14 Late gastrula 1.5 Embryonic shield 16 Head fold 17 Optic vesicles 18 Somite formation begins 19 Finfold 20 Optic cups 21 Auditory placodes 22 Lens forms 2.3 Otoliths 24 Pectoral fins 2.5 Retinal pigmentation 26 Heart pumping 27 Lens transparent 28 Mouth and anus open 29 Peritoneum pigmented .30 Yolk reduction 31 Prehatching 32 Hatching 33 Hatched, prcliorn larva MacFarlane and Norton: Nutntional dynamics of embryonic-stage Sebastes 275 Lipids were extracted from oocytes and embryos by the method of Bhgh and Dyer ( 1959). Total Upids were quantified by using thin layer chromatography with flame ionization detection (TLC-FID) by an latroscan TH-10 Mark III (MacFarlane et al., 1990; MacFarlane et al., 1993). Lipids were separated into steryl or wax ester, triacylglycerol, nonesterified fatty acid, cholesterol, and polar lipid classes on Chro- marods S-III in a solvent bath of hexane:ethyl ether:formic acid at a ratio of 246:54:0.09. Quantifi- cation of separated peaks by TLC-FID was accom- plished by comparing peak areas to external stan- dard curves for each lipid class. Cholesterol oleate, triolein, oleic acid, cholesterol, and phosphatidylcho- line, purchased from Sigma Chemical Co. (St. Louis, MO), were used as standards. Total protein was esti- mated with the Lowry method by using a bovine se- rum albumin standard (Lowry et al., 1951). Analysis of variance (ANOVA) was employed to assess variation in lipids and protein by embryo maturation stage (EMS), rockfish species, or popu- lation of shortbelly rockfish. Estimated concentra- tions of nutrients at parturition were obtained by linear regression. All analyses were performed with SAS statistical software (SAS Institute, Inc., 1989). Lipids and protein data are presented as concen- trations, mg/g wet weight. Changes in concentration during embryonic development represent absolute changes in the quantities of lipids and protein be- cause ovarian mass remained statistically constant through gestation for both species. ANOVA results of the relationship between GSI [gonadosomatic index = (ovary weight/ body weight) x 100] and EMS were P = 0.242 (F= 1.396, df=59) for yellowtail rockfish and P = 0.622 (F=0.243, df=202) for shortbelly rockfish. During gestation, changes in organic mass of ovaries were compensated by changes in water concentration. Regression equations for ovarian water concentration by EMS were Yellowtail rockfish ovarian water (mg/g) = 608.5-t-8.292(£;MS), [df=59, F=203.7, P<0.0001, r-=0.778] Shortbelly rockfish ovarian water (mg/g) "=655.2h-6.860(£MS). [df=172, F=782.6, P<0.0001, r^=0.821] Results and discussion There was a progressive decline in total lipid =and protein during embryogenesis in yellow- tail rockfish, Sebastes flavidus (Fig. 1). The o U concentration of total lipid decreased from a mean of 149.8 mg/g in unfertilized, late vitellogenic oocytes (EMS 1) to an estimated concentration of 48.4 mg/g for fully formed, hatched larvae (EMS 33) at partu- rition. Although no pregnant females were caught with embryos at EMS 33, the goodness-of-fit (r^ value) of the linear regression of lipid concentration on embryo maturation stage (Table 2) suggested that calculation of total lipid at parturition was valid (Table 3). Similarly, protein declined from 225.9 mg/g during EMS 1 to 52.1 mg/g at EMS 33 (Fig. 1; Table 3). These data indicate that yellowtail rockfish embryo- genesis consumed more protein than lipid for nutri- tional requirements because protein concentration declined 77% during development whereas total lip- ids decreased 68%. From an energetic perspective, however, lipid was predominantly utilized. Using values for physiologically available energy density of 39.6 kJ/g of lipid and 20.1 kJ/g of protein (Brett and Groves, 1979), we found that the 101.4 mg lipid/g embryo depleted during embryogenesis 3delded 4.02 kJ of energy, whereas the 173.8 mg protein/g embryo yielded 3.49 kJ. The predominance of lipid as an en- ergy source for viviparous reproductive development in yellowtail rockfish has been demonstrated previ- ously (MacFarlane et al., 1993; Norton and Mac- Farlane, 1995). The consumption of yolk proteins, particularly those with high molecular weights > 70 kDa, during embryonic maturation of yellowtail rock- Sebastes flavidus Tola! Protein Total Lipid / 1 6 11 16 21 26 31 Embryo maturation stage Figure 1 Total lipid and protein concentrations in oocytes and embryos by embryo maturation stage (EMS) in female yellowtail rockfish, Sebastes flavidus, from prefertilized oocytes (EMS 1) through the yolk-reduction stage (EMS 30). Data presented as means ± SE. Mean number (range) of females assayed at each EMS was 12 (9-17). See Table 1 for descriptions of EMS. 276 Fishery Bulletin 97(2), 1999 Table 2 Linear regression parameter estimates for mean protein and lipid concentrations in rockfi sh embryos in relation to embryo maturation stage (EMSi. Data are for vellowtail roc kfishiSff astes flavidus and three populations of s liortbelly rockfish iSebastes jordani) from Ascension Pioneer, and Bodega Canyons. ND = below detection (<0.3 ng); NS = = P > 0.05. TAG = triacylglycerols; NEFA = nonesterified fatty acids. CH = cholestero ; PL = polar lipids. Variable r^ Intercept ±SE P Slope ±SE P S. flavidus (/!=60) Total lipid 0.946 153.0 ± 7.6 <0.001 -3.17 ±0.44 <0.005 Total protein 0.993 231.3 ±4.6 <0.0001 -5.43 ±0.27 <0.001 S. jordani All canyons f.n = 212) Total lipid 0.884 89.4 ±3.8 <0.0001 -1.86 ±0.17 <0.0001 Total protein 0.895 167.8 ±6.0 <0.0001 -2.85 ±0.27 <0.0001 Esters 0.695 7.4 ± 0.5 <0.0001 -0.17 ±0.02 <0.0001 TAG 0.917 43.0 + 1.7 <0.0001 -0.98 ±0.08 <0.0001 NEFA 0.141 -O.ltO.l NS 0.01 ±0.00 NS CH 0.639 3.8 ±0.2 <0.0001 -0.05 ±0.01 <0.0001 PL 0.889 35.4 ± 1.4 <0.0001 -0.66 ± 0.06 <0.0001 Bodega Canyon in =49 Total lipid 0.699 81.0 ± 6.9 <0.0001 -1.54 ±0.32 <0.001 Total protein 0.676 174.1 ± 14.9 <0.0001 -3.17 ± 0.69 <0.001 Esters 0.750 7.3 ±0.6 <0.0001 -0.16 ± 0.03 <0,0001 TAG 0.704 37.4 ±3.5 <0.0001 -0.80 ±0.16 <0,0001 NEFA ND ND ND ND ND CH 0.503 3.4 ± 0.3 <0.0001 -0.04 ±0.01 <0.01 PL 0.665 32.8 ± 2.6 <0.0001 -0.54 ±0.12 <0.001 Pioneer Canyon i?i = 102) Total lipid 0.844 88.3 ±4.2 <0.0001 -1.97 ±0.18 <0.0001 Total protein 0.714 177.0 ±10.8 <0.0001 -3.29 ±0.46 <0.0001 Esters 0.766 7.4 ± 0.5 <0.0001 -0.18 ±0.02 <0.0001 TAG 0.835 44.6 ± 2.5 <0.0001 -1.12 ±0.11 <0.0001 NEFA 0.284 -0.2 ±0,1 NS 0.01 ±0.00 <0.01 CH 0.719 3.9 + 0.2 <0.0001 -0.06 ±0.01 <0.001 PL 0.838 32.7 ± 1.4 <0.0001 -0.62 ±0.06 <0.0001 Ascension Canyon i/i = 61) Total lipid 0.419 90.4 ± 8.0 <0.0001 -1.36 ± 0.41 <0.005 Total protein 0.451 165.1 ±11.8 <0.0001 -1.93 ±0.59 <0.005 Esters 0.227 7.2 ± 1.0 <0.0001 -0.11 ±0.05 <0.05 TAG 0.348 42.4 ± 4.6 <0.0001 -0.68 ±0.24 <0.01 NEFA 0.072 0.1 ± 0.1 NS 0.00 ± 0.00 NS CH 0.257 3.8 ±0.3 <0.0001 -0,03 ± 0.01 <0.05 PL 0.520 36.9 ± 2.6 <0.0001 -0.55 ± 0.14 <0.001 fish has also been reported (MacFarlane and Bow- ers, 1995). Calculations of lipid and protein utiliza- tion for embryogenesis in yellowtail rockfish repre- sent net consumption and conservative estimates because at least some maternal contributions to embryo development occur during gestation (Mac- Farlane and Bowers, 1995). Radio-labelled phos- phatidylcholine injected into the circulatory system of pregnant females resulted in radioactivity in em- bryos, indicating matrotrophy for this phospholipid. The extent to which matrotrophy exists for other lip- ids and protein is unknown. As in yellowtail rockfish, total lipid and protein concentrations declined linearly according to stage of development in embryos of shortbelly rockfish, Sebastes jordani (Fig. 2). Embryonic development consumed 55% of the pro- tein and 689f of the total lipid present at fertiliza- tion. Thus, both species of rockfish metabolized the same proportion of total lipids, whereas protein was conserved in shortbelly rockfish in relation to yel- lowtail rockfish embryogenesis. It should be noted, however, that yellowtail rockfish had significantly greater protein resources for embryogenesis at fer- MacFarlane and Norton: Nutritional dynamics of embryonic-stage Sebastes 277 Table 3 Estimated concentrations of lipids and protein in Sebastes flavidus and Sebastes jordani larvae at birth. S.jordani data are for all populations combined and for each of three populations at Bodega, Pioneer, and Ascension submarine canyons. Values presented as mean ± SD in mg/g wet weight. TAG = triacylglycerols; NEFA = nonesterified fatty acids; CH = cholesterol; PL = polar lipids. Sebastes flavidus Sebastes jordani Variable All Bodega Pioneer Ascension Protein 52.1 ±5,2 73.6 ± 6,4' 69.3 ±10.2 68,5 ± 6,3 101.3 ±9.1^ Total lipid 48.4 + 8,5 28.1 ± 3.72 29.4 ±4.7-3 23.3 ± 2.7 45.7 ± 6.4^ Esters — 1.9 ± 0.4 1.9 ± 0.4'' 1.4 ± 0.3 3.8 ± 0.7-' TAG — 10.5 ±2.0 11.0 ±2.4-' 7.6 ± 1.6 20.1 ±3.7^ NEFA — 0.1 ±0.1 — 0.2 ± 0.0 0.0 ±0.1 CH — 2.1 ±0.1 2.0 ± 0.2 2.0 ± 0.1 2.9 ± 0.2^ PL — 13.6 ± 1.4 14,5 ± 1,8^ 12.1 ±0,9 18.9 ±2.1-' ' Significantly greater than that for S. flavidus iP<0.0001 1. - Significantly less than that for S. flavidus iP<0.01 1. ■* Greater than that for S, jordani at Pioneer Canyon (P<0.001 1, ■• Greater than that for S jordani at Bodega and Pioneer Canyons iP<0 0001). tilization than did shortbelly rockfish, 225,9 mg/g compared with 165,0 mg/g (Table 2), At birth, the estimated concentration of protein in shortbellly rockfish larvae was signifi- cantly greater than that in newborn yellow- tail rockfish (P<0,0001); however, the total lipid concentration was less (P<0,01) (Table 3), That shortbelly rockfish have less lipid at parturition suggests they have a greater ur- gency for feeding than yellowtail rockfish soon after birth, or else catabolism of tissue protein would be required for maintenance, reducing the probability of survival. Energetically, lipid was the predominant fuel for embryo development for shortbelly rockfish, as for yellowtail rockfish. The 59.4 mg lipid/g embryo metabolized between fer- tilization and parturition represented 2,35 kJ/g, and the 91,3 mg protein/g embryo rep- resented 1,84 kJ/g, Total energy consumed for embryogenesis was significantly greater in yellowtail rock- fish than in shortbelly rockfish (P<0,0001), The production of a gram of fully formed, hatched, preborn larval yellowtail rockfish required the expenditure of 7,51 kJ of energy. For shortbelly rockfish, the equivalent energetic cost was 4,19 kJ/g of EMS-33 larvae. Because the aver- age weight of ovaries at late stages of gestation was 204,9 ± 16,9 g (n = 17) in yellowtail rockfish but only 43.4 ± 2.8 g {n-21) in shortbelly rockfish, require- - ments for maternally supplied nutrients were much greater for yellowtail rockfish. For comparison, the 250 200 150 Sebastes jordani Total Protein Total Lipid 1 6 11 16 21 26 31 Embryo maturation stage Figure 2 Total lipid and protein concentrations in oocytes and embryos in fe- male shortbelly roektish, Sebastes jordani . from prefertilized oocytes (EMS 1) through the hatched preborn larval stage (EMS 33). Data I means ± SE) from shortbelly rockfish from populations at all three submarine canyons are shown. Mean number (range) of females as- sayed at eachEMS was 8 (1-21). See Table 1 for description of EMS. mean standard length and total weight for female yellowtail rockfish was 36 ± 0.5 cm and 1304 ± 55 g versus 21 ± 0,2 cm and 150 ± 4 g for the shortbelly rockfish assessed in this study. The significant difference in energy required to produce a gram of larvae in the hatched, preborn stage in the two rockfish species may relate to differences 278 Fishery Bulletin 97(2), 1999 in reproductive strategy. Yellowtail rockfish produce about 700,000 eggs/female (Eldridge and Jarvis, 1995) whereas shortbelly rock- fish fecundity is about 16,500 eggs/female (EldridgeM. Also, yellowtail rockfish eggs and hatched larvae are considerably smaller than those of shortbelly rockfish (Matarese et al., 1989; Eldridge^). Greater energy consump- tion for production of yellowtail rockfish lar- vae may be a consequence of greater energy required for the synthesis of more larvae, their organelles, and subsequent cellular dif- ferentiation. The relative importance of lipids and pro- tein as sources of nutrition during embryonic and larval development in oviparous fishes has been presented in other reports. In the marine teleost, Spai'us aurata, free amino acids were the main source of energy during the relatively short embryonic period (ca. 51- 58 h postfertilization) whereas lipids and protein did not supply substantial energy until after hatching (Ronnestadet al., 1994). Winter flounder iPseiidopleuronectes ameri- canus), by contrast, used small amounts of carbohy- drate first, then protein until hatching (Cetta and Capuzzo, 1982). Lipid depletion did not occur until after hatching. Protein concentrations were stable, whereas lipids were catabolized in eggs and newly hatched larvae of Senegal sole (Solea senegalensis) (Mourente and Vasquez, 1996). According to these researchers, this pattern of lipid metabolism is char- acteristic of temperate marine fishes with eggs con- taining oil globules, high lipid content, and short de- velopmental periods. The rockfish species in the present study are temperate marine teleosts that contain oil globules and high lipid levels. However, they are viviparous and have relatively lengthy de- velopment, yet catabolize both protein and lipids for energy-consuming processes, thus emphasizing the great variability and lack of generality in the sources and patterns of nutrient use for embryonic and lar- val development in fishes. Total lipid concentrations can be considered a mea- sure of physiological condition of larvae (Ferron and Leggett, 1994), and thus an indicator of potential reproductive success, but not all types, or classes, of lipids are equivalent with respect to metabolic avail- ability or in vivo energy yield. Therefore, fraction- ation of total lipids into classes representative of energy-yielding and structural functions provides Sebastes jordani Embryo maturation stage Figure 3 Changes in lipid classes during embryonic maturation in shortbelly rockfish, Sebastes jordani. Data (means ± SEl are from populations at all three submarine canyons. Number of females assayed for each EMS is same as in Figure 2. See Table 1 for description of EMS. ' Eldrige. M. B. 1997. ies Science Center. 94720. Unpubl. data. Tiburon Laboratory, Southwest Fisher- 3150 Paradise Drive, Tiburon. CA greater knowledge of the amount of energy available to sustain growth and survival once the larvae are released into the environment. Total lipid from shortbelly rockfish embryos was fractionated into five lipid classes: steryl/wax esters (esters), triacylglycerols (TAGs), nonesterified fatty acids (NEFAs), cholesterol (CH), and polar lipids (PLs). The separations were not performed on yel- lowtail rockfish embryos owing to loss of samples before chromatographic analysis. Triacylglycerols (TAGs) and PLs were the predomi- nant lipids in all stages of embryonic maturation and showed the greatest declines during development (Fig. 3). Cholesterol (CH) and esters were found in much lower concentrations and declined slightly, but significantly, during embryogenesis (Fig. 3; Table 2). Nonesterified fatty acids (NEFAs) were found in very low concentrations and did not vary significantly by stage of maturation (Table 2). The depletion of TAG and PL from fertilization to birth indicates the oxidation of these lipid classes for energy-requiring embryogenic processes. The use of TAG for energy storage and fuel during embry- onic and larval development is well known in fishes (Boulekbache, 1981; Vetter et al., 1983; Tocher and Sargent, 1984; Eraser, 1989). Polar lipids, consist- ing mainly of phospholipids, such as phosphatidyl- choline and phosphatidylethanolamine, although used as fundamental structural units of all biologi- cal membranes, have been shown to be a significant source of energy also for embryonic development in MacFarlane and Norton; Nutritional dynamics of embryonic-stage Sebastes 279 other fishes (Sargent, 1995). The relative contribu- tion of TAG and PL to the energy demands of embry- onic development varies by species. Whereas in cod iGadusmorhua) iFraseretal., 1988; Finn etal., 1995) and Atlantic herring iClupea harengus) (Tocher et al., 1985) PL was the major source of energy, in Senegal sole (Mourente and Vasquez, 1996) and gilthead sea bream ( Ronnestad et al., 1994), TAG was primarily catabolized. In Atlantic salmon (Salmo salar), the PL phosphatidylcholine and TAG were used equally to meet energy requirements (Cowey etal., 1985). The changes in lipid composition during embryo- genesis in yellowtail and shortbelly rockfishes de- scribed here represent the first data of this kind for rockfishes or any viviparous marine teleost. The use of PL as a primary source of energy may be consis- tent with results from previous research revealing that phosphatidylcholine was supplied to develop- ing embryos of yellowtail rockfish from maternal sources both before and during development (MacFarlane and Bowers, 1995). Perhaps the facil- ity to supply phospholipids matrotrophically en- hances their use as energy-yielding substrates. High concentrations of TAG in prefertilized eggs (EMS 1) and its importance as an energy source are not un- expected because both species feed heavily on eu- phausiids (Brodeur and Pearcy, 1984), which contain large amounts of TAG (Fricke and Oehlenschlager, 1982). Lipid class composition has been promoted as a measure of nutritional condition for larvae of a vari- ety of oviparous marine fish, including Atlantic cod (Lochmann et al., 1995), Atlantic herring (Fraser et al., 1987), and anchovy Engraul is mordax (Hakanson, 1989). Depending upon the species, the quantities of TAG or PL, or both, related to the probability for survival. Lochmann et al. ( 1995) suggested that sur- vival was related to a condition index based on a dis- criminant function of TAG, PL, and defatted dry weight in laboratory and field-collected cod larvae. Triacylglycerol (TAG) was regarded as the critical variable in condition for herring (Fraser et al., 1987) and anchovies (Hakanson, 1989). It is reasonable to apply this concept to the assess- ment of larvae from viviparous species as well. The nutritional status of larvae just prior to parturition would represent their condition at the start of life in the marine environment and would be related to their ability to survive until adequate feeding occurs. The presence of an oil globule in rockfish larvae during the early days after birth argues for the importance of lipids for energy. For rockfish, the amounts of TAG ■ at birth should be related to survival potential be- cause this lipid class is the predominant component in oil globules of most fishes (Henderson and Tocher, 1987; Heming and Buddington, 1988) and has con- sistently been shown to correlate with physiological condition (Fraser, 1989). We analyzed embryos in pregnant female short- belly rockfish from populations living in the proxim- ity of three submarine canyons off the central Cali- fornia coast. The purpose was to determine the vari- ability in metabolism of lipid classes for specific popu- lations and the estimated concentrations of each lipid class at birth, thus providing a measure of nutritional status and relative probability for survival. When lipid class concentrations were assessed in embryos from the separate populations, there were differences in the amounts and rates of depletion during intraovarian development (Table 2). Two-way ANOVA determined there were significant differences among populations at the three canyons, despite being lo- cated within about a one-degree latitude span off the central California coast. All lipid classes, except NEFA, varied significantly by EMS (P<0.0001 ), popu- lation (P<0.0001), and the interaction of population and EMS (P<0.05). The changes in concentrations of lipid classes were linearly related to EMS for embryos from each popu- lation (Table 2). Parameter estimates from linear regressions of each lipid class by stage of embryonic maturation revealed that rates of metabolism dif- fered among the populations (Table 2). The greatest catabolic rates of TAG and PL were found in embryos from the Pioneer Canyon population. Rates of pro- tein depletion were also greater in embryos from Pio- neer Canyon and lowest in those from Ascension Canyon. In conjunction with initial concentrations of pro- tein and lipid classes at fertilization, the rates of nutrient metabolism can be used to estimate concen- trations at parturition (EMS 33). The greater rates of nutrient utilization for embryos in females from the population at Pioneer Canyon resulted in the low- est levels of nutritional stores at birth of the three populations evaluated (Table 3). Conversely, newborn larvae from females at Ascension Canyon contained the greatest concentrations of lipids and protein, suggesting that they have a greater likelihood of sur- vival until suitable feeding conditions exist. The differential status of nutrient levels in larvae at birth may contribute to differences in survival rate among the populations and, thus, influence the di- versity of the species year class. Parturition occurs during the winter when oceanic conditions prevail along the California coast. This is the period of low- est primary and secondary productivity annually (Ainley, 1990). Until marine conditions become fa- vorable for biological productivity, usually in April 280 Fishery Bulletin 97(2), 1999 when seasonal upwelling and increasing solar radia- tion stimulate phytoplankton growth, larvae must subsist on the combined resources of endogenous re- serves and limited prey abundance. During the days following birth, the quantity of lipid reserves may influence the length of time larvae can persist until adequate forage is available to sustain survival and growth. Starvation has been considered a factor in recruitment variability in rockfish species ( Moser and Boehlert, 1991). Larvae from populations with lesser energy reserves may experience greater mortality, thereby reducing their representation in the year class for the species along the entire coast. The use of lipid and protein data as an indicator of nutritional condition at birth for rockfish in the ge- nus Sebastes marks an extension of such data to vi- viparous fishes for the first time. Previously, lipid class analyses have been employed for condition as- sessments only in oviparous fish embryos (Tocher et al., 1985; Fraser et al., 1988) and larvae ( Fraser, 1989; Hakanson, 1989; Lochmann et al., 1995). Acknowledgments We wish to express our appreciation to Maxwell Eldridge, Brian Jarvis, and Michael Bowers for their assistance with collecting the fish used in this re- search. We also thank Richard Powers, owner and operator of the FV New Sea Angler, as well as the officers and crew of the NOAA ship David Starr Jor- dan for their invaluable expertise and effort during marine operations. Literature cited Ainley, D. G. 1990. Seasonal and annual pattern.'; in the marine envi- ronment near the Farallones. In D. G. Ainley and R. J. Boekelhcide I eds. ), Seabirds of the Farallon Islands, p. 23- .50. Stanford Univ. Press, Stanford, CA. Ainley, D. G., R. H. Parrish, W. H. Lenarz, and W. J. Sydeman. 1993. Oceanic factors influencing distribution of young rock- fish (Sebastes) in central California; a predator's perspective. Calif. Coop. Oceanic Fish. Invest. Rep, 34:1.33-1:39, Bakun, A. 1975. Daily and weekly upwelling indices, west coast of North America, 1967-73, US, Dep, Commer NOAA Tech, Rep NMFS SSRF-693, 114 p, Blaxter, J. H. S. 1969. Development: eggs and lafvae. In W. S. Hoar and D J. Randall (eds, I, F'ish physiology, vol. III: Reproduction and growth, p. 177-2.52, Academic Press. New York. NY, Blaxter, J. H. S., and G. Hempel. 196.3. The influence of egg size on herring larvae iCliipea harengus). J, Cons, Int, Explor. Mer 28:211-240. Bligh, E. G., and G. Dyer. 1959. A rapid method of total lipid extraction and purification. Can. J, Biochem. Physiol, 37:911-917, Boulekbache, H. 1981. Energy metabolism in fish development. Am, Zool. 21:377-389, Brett, J. R., and T. D. D. Groves. 1979. Physiological energetics. In W. S. Hoar, D, J. Randall, and J. R. Brett (eds.) Fish physiology, vol, VIII: Bioenergetics and growth, p, 280-352. Academic Press, New York, NY, Brodeur, R. D., and W. G. Pearcy. 1984. Food habits and dietary overlap of some shelf rock- fishes (genus Sebastes) from the northeastern Pacific Ocean, Fish, Bull, 82:269-293. Buckley, L. J. 1984. RNA-DNA ratio: an index of larval fish growth in the sea. Mar Biol, 80:291-298, Cetta, C. M., and J. M. Capuzzo. 1982. Physiological and biochemical aspects of embryonic and larval development of the winter flounder (Pseudo- pleuronectes amcncanus). Mar Biol, 71:327-337, Clarke, M. E., C. Calvi, M. Domeier, M. Edmonds, and P. J. Walsh. 1992. of nutrition and temperature on metabolic enzymes activities in larval and juvenile red drum, Sciaenops ocellatus, and lane snapper, Liitjanus synagris. Mar. Biol, 112:31-36, Cowey, C. B., J. G. Bell, D. Knox, A. Fraser, and A. Youngson. 1985. Lipids and antioxidant systems in developing eggs of salmon tSahno salar). Lipids 20:567-572, Eldridge, M. B., and B. J. Jarvis. 1995, Temporal and spatial variation in fecundity of yel- lowtail rockfish. Trans, Am, Fish, Soc, 124:16-25, Ferron, A., and W. C. Leggett. 1994. An appraisal of condition measures for fish larvae. Adv. Mar, Biol, 30:217-303, Finn, R. N., J. R. Henderson, and H. J. Fyhn. 1 995. Physiological energetics of developing embryos and yolk- sac larvae of Atlantic cod iGadus morhua). II. Lipid metabo- lism and enthalpy balance. Mar Biol. 124:371-379, Fraser, A. J. 1989. Triacylglycerol content as a condition index for fish. bivalve, and crustacean larvae. Can. J, Fish, Aquat. Sci, 46:1868-1873. Fraser, A. J., J. C. Gamble, and J. R. Sargent. 1988. Changes in lipid content, lipid class composition, and fatty acid composition of developing eggs and unfed larvae of cod iGadus morhua). Mar Biol. 99:307-313. Fraser, A. J., J. R. Sargent, J. C. Gamble, and P. MacLachlan. 1987. Lipid class and fatty acid composition as indicators of the nutritional condition of larval Atlantic herring. Am. Fish. Soc, Symp, 2:129-143, Fricke, H. S. G., and J. Oehlenschlager. 1982. Separation of lipids from Antarctic krill (Euphausia superba Danal by isocratic high-performance liquid chro- matography on silica using a flow-program, J, Chroma- tography 252:331-334. Hakanson, J. L. 1989. Analysis of lipid components for determining the con- dition of anchovy larvae, Engraulis mordax. Mar, Biol, 102:143-151, Heming, T. A., and R. K. Buddington. 1988. Yolk absorption in embryonic and larval fishes In W, S. Hoar, and D. J. Randall (eds.) Fish physiology, vol. XI B, p. 407-446. Academic Press. San Diego, CA. MacFarlane and Norton: Nutritional dynamics of embryonic-stage Sebastes Henderson, R. J. and D. R. Tocher. 1987. The lipid composition and biochemistry of freshwa- ter fish. Prog. Lipid Res. 26:281-347. Lochmann, S. E., G. L. Maillet, K. T. Frank, and C. T. Taggart. 1995. Lipid class composition as a measure of nutritional con- dition in individual larval Atlantic cod (Gadus morhua). Can. J. Fish. Aquat. Sci. 52:1294-1306. Lowry, O. H., N. J. Rosebrough, A. L. Farr, and R. J. Randall. 1951. Protein measurement with the folin phenol reagent. J. Biol. Chem. 193:265-275. MacFarlane, R. B., and M. J. Bowers. 1995. Matrotrophic viviparity in the yellowtail rockfish Sebastes flavidus. J. Exp. Biol. 198:1197-1206. MacFarlane, R. B., H. R. Harvey, M. J. Bowers, and J. S. Patton. 1990. Serum lipoproteins in striped bass (Morone saxatilis): effects of starvation. Can. J. Fish. Aquat. Sci. 47:739-745. MacFarlane, R. B., E. C. Norton, and M. J. Bowers. 1993. Lipid dynamics in relation to the annual reproduc- tive cycle in yellowtail rockfish tSebastes flandust. Can. J. Fish. Aquat. Sci. 50:391-401. Matarese, A. C, A. W. Kendall Jr., D. M. Blood, and B. M. Vinter. 1989. Laboratory guide to early life history stages of north- east Pacific fishes. U.S. Dep. Commer NOAA Tech. Rep. NMFS 80. 652 p. May, R. C. 1974. Larval mortality in marine fishes and the critical period concept. /;; J. H. S. Blaxter led.t, The early life history of fish. p. 3-19. Springer- Verlag, New York, NY. Moser, H. G., and G. W. Boehlert. 1991. Ecology of pelagic larvae and juveniles of the genus Sebastes. Environ. Biol. Fishes 30:203-224. Mourente, G., and R. Vasquez. 1996. Changes in the content of total lipid, lipid classes and their fatty acids of developing eggs and unfed larvae of the Senegal sole, Solea senegalensis Kaup. Fish Physiol. Biochem. 15:221-235. Norton, E. C, and R. B. MacFarlane. 1995. Nutntional dynamics of reproduction in viviparous yel- lowtail rockfish, Sebastes flavidus. Fish. Bull. 93:299-307. Ralston, S., and D. F. Howard. 1995. On the development of year-class strength and co- hort variability in two northern California rockfishes. Fish. Bull. 93: 710-720. Rlssik, D., and I. M. Suthers. 1996. Feeding in a larval fish assemblage: the nutritional significance of an estuarine plume front. Mar. Biol. 125:233-240. Rennestad, I., W. M. Koven, A. Tandler, M. Harel, and H. J. Fyhn. 1994. Energy metabolism during development of eggs and larvae of gilthead sea bream iSparus aurata). Mar Biol. 120:187-196. Sargent, J. R. 1995. Origins and functions of egg lipids: nutritional implications. In N. R. Bromage and R. J. Roberts leds.), Broodstock management and egg and larval quality, p.353- 372. Blackwell Science, London. SAS Institute, Inc. 1989. SAS/STAT user's guide, 4'hed., vol. 2. version 6. SAS Institute, Cary, NC, 846 p. Shepherd, J. G., and D. H. Gushing. 1980. A mechanism for density-dependent sun'ival of lar- val fish as the basis of a stock-recruitment relationship. J. Cons. Int. Explor Mer 39:160-167. Theilacker, G. H. 1978. Effect of starvation on the histological and morpho- logical characteristics of jack mackerel, Trachurus symmetricus, larvae. Fish. Bull. 76:403-414. Tocher, D. R., A. J. Eraser, J. R. Sargent, and J. C. Gamble. 1985. Lipid class composition during embryonic and early larval development in Atlantic herring iClupea harengus, L.). Lipids 20:84-89. Tocher, D. R., and J. R. Sargent. 1984. Analyses of lipids and fatty acids in ripe roes of some northwest European marine fish. Lipids 9:492-499. Vetter, R. D., R. E. Hodson, and C. Arnold. 1983. Energy metabolism in a rapidly developing marine fish egg. the red drum iSciaenops ocellata I. Can. J. Fish. Aquat. Sci. 40:627-634. Wyllie Echeverria, T. 1987. Thirty-four species of California rockfishes: maturity and seasonality of reproduction. Fish. Bull. 85:229-250. Yamada, J., and M. Kusakari. 1991. Staging and the time course of embryonic develop- ment in kurosoi. Sebastes schlegeli. Environ. Biol, Fishes 30:147-153. 282 Abstract.— Respiration rates of non- feeding adult menhaden induced to swim against currents of various speeds in a large circular flume at 10', 15 . and 20 C were measured in order to quan- tify the cost of swimming separately from total metabolic expenditure dur- ing filter feeding. Standard metabolic rates of 0.040, 0.073, and 0.087 mg O,^ / (g wet wt ■ h) at 10°C, 15°C, and 20X were estimated by extrapolation of the relationship of swimming speed and metabolic rate to zero swimming speed. We determined that when menhaden filter-feed at 200. at the preferred swimming speed of 41.3 cm/s, filtering and specific dynamic action (SDA) ac- count for 59'7c of total energetic expen- ditures. The cost of locomotion was only about 23*7^ of the total expenditure. Our results are compared with routine oxy- gen consumption rates of larval and juvenile menhaden as a function of tem- perature and with extensive metabolic data for sockeye salmon. Metabolic rate in relation to temperature and swimming speed, and the cost of filter feeding in Atlantic menhaden, Brevoortia tyrannus William K. Macy ill Ann G. Durbin Edward G. Durbin Graduate School of Oceanography University of Rhode Island Soutfi Ferry Road Narragansett, Rhode Island 02882 E-mail address (for W K Macy) wkmacyagsosunl gsouri edu Manuscript accepted 23 June 1998. Fish. Bull. 97(2):282-293 (1999). The Atlantic menhaden, Brevoortia tyrannus, is a pelagic filter-feeding clupeoid inhabiting coastal waters of North America from Maine to Florida. Menhaden adopt filter feed- ing in early juvenile life ( Peck, 1893; June and Carlson, 1971; Friedland et al., 1984 ). Adults are obligate fil- ter feeders and consume phyto- plankton, zooplankton, and detritus larger than about 13 ).im (Durbin and Durbin, 1975). Menhaden are size selective in their feeding, but the basis of selection is passive, re- flecting the size distribution of par- ticulate material in the water rela- tive to the "pore size" of the gill rak- ers; they do not actively pursue in- dividual prey. Energy expenditures during "ram" filter feeding (analo- gous to towing a plankton net) are high (Durbin et al., 1981; Sanderson and Cech, 1992), probably due to the hydrodynaniic drag of the filtering apparatus. However, menhaden compensate for the high cost of fil- ter feeding by regulating their swimming speed to maximize the ratio of net energy gain to expendi- ture (Durbin and Durbin, 1983), as predicted by Ware ( 1975). Our study was undertaken to partition the previously determined energetic cost of filter feeding (Durbin and Durbin, 1983) into two components: the cost of locomotion alone and the cost of filter feeding. This division was accomplished by measuring respiration rates of nonfeeding fish forced to swim against currents of known speed and by comparing these respiration rates with those of feeding fish. Materials and methods Live Atlantic menhaden, Brevoortia tyrannis, were collected from a com- mercial fish trap located in the lower West Passage of Narragansett Bay on 22 and 25 September 1989. The fish were removed from the trap with dip nets and placed in 114-L seawater-fiUed polyethylene gar- bage cans for transport to the Narragansett Bay Campus. Menha- den were randomly divided and placed with three open-flow circu- lar fiberglass tanks, 1.7 m in dia- meter and 0.5 m deep, supplied with sand-filtered seawater. Acclimation to the three test temperatures ( 10°, 15", and 20 C) began 5 October 1989, starting from the ambient temperature, 18°C, with a rate of change of about l°C/d. Ambient sa- linities varied between 28 and 30 ppt. The fish were considered accli- mated after maintenance for 30 days at the target temperature. Menhaden were fed salmon starter mash (Ziegler Bros., Inc.) at a rate of about 4.5'7f of dry food:live Macy et al,: Metabolic rate of Brevoortia tyrannus 283 Table 1 Vital statistics for the experimental fish are given in the upper panel. Relationships between respiration and swimming speed (S, cm/s or body lengths [BL/s)) are given in the lower panel. T = temperature an fed for 24 h, and FF = filter feeding. rate ( R, mg Oj/Cg wet wt ■ h )) d FL = fork length, NF = not Trc) No. offish in sample FL(cm) Wet wt (g) Dry wt (g) 10 13 26.0 ± 0.9 327 ± 45 99.2 ± 15.6 15 13 25.5 ± 0.6 303 ± 38 92.1 ± 15.2 20 10 25.5 ± 1.2 283 ± 40 86.4 ± 12.3 T{°C) Behavior Regression equation r2 Source 10 NF log,(, R = 0.0141 S (cm/s) - 1.402 logi„ R = 0.3655 S (BL/s) - 1.598 0.99 This study 15 NF log,,, R = 0.0095 S (cm/s) - 1.139 logi„ R = 0.2419 S (BL/s) - 1.861 0.90 20 NF log,„ R = 0.0086 S (cm/s) - 1.061 logjo R = 0.2187 S (BL/s) - 1.939 0.96 20 FF log,„ R = 0.0271 S (cm/s) - 1.446 0.84 Durbin et al. (19811 weight/d. The acclimation tanks were cleaned daily. During the holding and experimental periods, a 12 h:12 h lightidark photoperiod was provided by over- head fluorescent lights, and sufficient current was provided to cause the fish to orient head into the cur- rent. Experiments took place 3-14 December 1989. Vital statistics for the fish are given in the top sec- tion of Table 1. Respiration measurements were made on a small school of menhaden because studies of schooling fishes have shown that respiration rates are higher and growth rates lower in isolated fish than in those kept in groups (Skazhina, 1975; Kanda and Itazawa, 1978; see also Ross and Backman, 1992). Experi- ments were begun with the 15°C trials on 3 and 4 Dec, followed by the 10°C trials on 7 and 8 Dec. The 20°C trials were conducted on 14 Dec. Thirteen fish were used for the 10°C and 15°C experiments. For the 20°C trials only 10 fish were used. Digestion is rapid in menhaden (Durbin and Durbin, 1981), and metabolism quickly returns to preceding levels after a meal (Durbin et al., 1981). Thus a starvation pe- riod of 24 h was considered sufficient to remove the effect of the previous meal on metabolic-rate mea- surements. Handling and transfer protocols were similarly chosen to reduce bias due to stress-induced factors, such as those identified by Waring et al. (1992) in flounder and salmon. A sealed, toroidal fiberglass flume tank (Fig. lA), modified from Hettler ( 1976), was used as a respirom- eter. The flume (2.75 m inside diameter, 0.51 m depth, and 0.39 m wide annular swimming channel) was large enough to hold a small school of adult menha- den. The flume was painted with a white, nontoxic, polyester gel coat. Black radial stripes 5.1 cm wide, spaced 21.8 cm apart (at the inside diameter of the flume), were painted on the walls and bottom of the flume, to provide reference marks for later determi- nation of swimming speeds and to provide visual cues for the fish to sense displacement by the flow. To pro- duce the desired flow rates, a high-pressure, 3.7-kW, multi-stage electric pump (Fig. lA) was connected to coils of polyethylene tubing (6 mm inside diam- eter) arrayed along the walls of the flume (Fig. IB). Discharge holes were punched at an angle of about 45° downstream and 20° downward from the hori- zontal with a hypodermic needle along the length of tubing to form small jets creating a counterclockwise flow in the flume. A valve placed at the pump outlet (Fig. lA) controlled the speed of water in the flume. Return water to the pump flowed through a drain line located at the bottom of the tank and thence through a bank of 4 fiber-wound cartridge filters (3 |am). An integral reservoir (12. 9-L capacity) was used to replenish sample water removed during experi- ments. Cooling coils were installed along the inner wall of the flume and around the pump housing (Fig. IB) to remove heat generated by the pump and to maintain a constant temperature during trials. Wa- ter temperature was monitored with a YSI Model 2100 Tele-thermometer. Flow rates within the flume were measured with a General Oceanics Inc. flowmeter connected to a proprietary digital display. The transducer head was 284 Fishery Bulletin 97(2), 1999 2 75m inside discharge manifolds filter bank pressure gauge flow adjusting valves cooling coil pump flow meter transducer B inlet discharge lines 0.51m r ^ sampling port 1 , 606 . 600 + ' 73.0 + <^ ¥ 73 66 H 65 H .8 . t .9 • « 62.8 + 64.0 + 63.4 + \ \\ ' 4 ' * 0.39m flow balancing valves Figure 1 (Al Schematic view of the toroidal flume respirometer. The direction of flow is indicated by arrows within lank. (B) Cross-sectional view of flume tank show- ing arrangement of water and coolant flow. A typical cross section of the gen- erated flow field is also shown. Speeds are cm/s. The grid spacing is 10.8 cm vertically by 8.3 cm horizontally. suspended in the flume by a stain- less steel rod (Fig. IB). During experi- ments the flowmeter head was fixed radially in the middle of the flume, near the top of the water column. However, sensor depth and radial position could be varied to measure flow rate in different regions of the flume. By varying the number of dis- charge holes across the vertical ar- ray of inlet coils and by regulating the flow to the inner and outer inlet coils, it was possible to produce a relatively uniform cross-sectional flow field. Typical flow rates (cm/s) determined at maximum pump pressure (1.38 x lO'^kPa ), on a grid 10.8 cm vertically by 8.3 cm horizontally measured ver- tically from the flume bottom, are shown in Figure IB. Preliminary tri- als established that the relationship between pressure of the inflowing water from the pump and the flow rate generated inside the flume was linear up to the maximum pump pres- sure, 1.38 X lO'^ kPa, corresponding to a flume speed of 64 cm/s. Blank trials without fish demonstrated that the system was airtight and that changes in oxygen concentration due to microbial respiration were negligible over periods comparable to those in the experiments. Each experiment comprised a se- ries of trials over a 2-d period. Men- haden were anesthetized with MS- 222 and transferred to the flume 24 h before an experiment to permit re- covery from handling stress (Barton and Schreck, 1987; Waring et al., 1992). The menhaden remained in the flume until the conclusion of the experiment and were fed in the late afternoon of each day. In the morn- ing, fecal material and debris was si- phoned off the bottom of the flume, the bottom and interior walls of the flume were scrubbed with a polyeth- ylene scouring pad, and the flume was flushed with filtered seawater. All experiments were conducted dur- ing daylight in the afternoon, 24 h after the last feeding. The fish were accustomed to this procedure and quietly circuited the flume. A clear Macy et al.: Metabolic rate of Brevoortta tyrannus 285 Plexiglas lid, composed of interlocking sections, was then carefully fitted over the surface of the tank to provide an airtight seal. During respi- ration trials, the flume was operated as a closed system; between trials and at night, however, the flume received a continuous inflow of tem- perature-controlled filtered seawater, with a flow rate of about 6 cm/s to orient the fish. During each 2-d experiment a series of trials was carried out at flume speeds ranging from 12-13 (minimum) to 60-63 (maximum) cm/s. Between four and six trials were conducted per day, with each trial lasting 1 h. Test speeds were selected randomly, and at the completion of a trial the next test speed was set, followed by a 30-min interval during which flow rate became stable again and fish were able to adjust to the new speed. An interval of about 1 h was used to reoxygenate the flume when necessary. Test current speeds were set by regulating the dis- charge pressure of the circulating pump (Fig. lA) by using the predetermined pressure to flow-rate calibration curve. Although we at- tempted to use the same test current speeds at all three temperatures, minor inaccuracies in the calibration curve and gauge readings resulted in only approximately equal test speeds (Fig. 2). Fish behavior was observed continuously during experiments and recorded with a mono- chrome MTI Dage M65 video camera suspended directly above the center of the flume and con- nected to a VHS format video recorder. Swim- ming speed in relation to the bottom (ground speed, GS) was determined from the video re- cordings at 3-min intervals, by marking the position of each fish on the video monitor screen with a grease pencil and noting its subsequent location 10 s later. The number of reference marks crossed by each fish during this interval was then multiplied by the distance between marks (21.8 cm) to obtain ground speed. Fish moving upstream were assigned positive values, whereas those swimming downstream were scored negatively. Actual swimming speed ( SS ) through the water was then computed as the algebraic sum of the ground speed (speed offish in relation to the bot- tom) and the fiow rate in the flume. Mean swimming speed during a trial was computed from the indi- vidual swimming speeds, irrespective of direction. During experiments a YSI Model 51B dissolved oxy- gen meter was used to monitor oxygen levels in real time; oxygen content of the water was not permitted to decline below 607f of saturation. At 15-min intervals during a trial, duplicate water samples were siphoned fi-om the collection port into 300-mL biological oxygen 70 -| 60- 50- 40- 30 20 10 1 o ■d 10°C • SS = 9.49 X F + 8.57. r = 0.93 O GS= -0.45 X F + 8.53, r^ = 0.03 •^^i^^ii— cNr^ 15°C • SS= 9.77xF + 603. r =099 O GS = -0.33 X F + 6.65. r = 0.14 20°C • SS= 1.01 xF-^ 2 07. r" = 0 99 O GS = -0 49 X F -^ 6.73, r^ = 0.30 -| r 0 10 20 30 40 50 60 70 Flume Speed, cm/s Figure 2 Ground speeds (GS) and true swimming speeds (SSl of Atlantic menhaden, Brevoortia tyrannus, swimming against water cur- rents in a toroidal flume respirometer at 10', 15", and 20°C. Lower ground speed error bars are not shown. demand (BOD) bottles that were allowed to overflow twice their volume and that were then fixed for deter- mination of oxygen content by Winkler titration ( Carritt and Carpenter, 1966; Strickland and Parsons, 1972). Analyses were performed on 5-6 50 niL subsamples withdrawn fi-om the bottles with a volumetric pipette. Dissolved oxygen declined linearly over time (overall mean r^ ± 1 SD for all regressions=0.987 ±0.018). Meta- bolic rates were therefore assumed to be unaffected by declining oxygen content during the trials. Menhaden respiration rates were computed fi-om the change in dissolved oxygen content over time as determined by least squares linear regression, corrected for the wet weight of the fish and volume of the flume. 286 Fishery Bulletin 97(2), 1999 Results The menhaden formed a loose school in the upper central portion of the flume. The fish usually oriented head into the current and moved slowly upstream (positive ground speed). Occasionally a fish would break away and swim downstream with the flow, but within a few seconds reoriented itself to swim against the flow. These brief periods of downstream swim- ming occurred most frequently at slow current speeds and at the warmest temperature, 20°C. The fre- quency of occurrence of downstream coasting or swimming during respiration trials was as follows: 0 of 11 trials at 10°C: 2 of 11 at 15°C (2.2'7f of obser- vations during those 2 trials); and 6 of 10 at 20°C (6.4% of observations in those 6 trials). When cor- rected for current speed, swimming speeds of fish moving downstream were comparable to those offish moving upstream. Overall, the incidence of downstream swimming did not significantly affect the results. In our study, menhaden swimming into a water current maintained a nearly constant ground speed at 10°, 15°, and 20°C (6.8, 5.3, and 4.7 cm/s respec- tively, or about 0.2 BL/s) (Fig. 2), whereas their true swimming speeds ranged from 20.5 to 64.6 cm/s (10°C), 19.2 to 65.7 cm/s (15 C), and 15.5 to 66.0 cnVs (20°C). The mean ground speed was not signifi- cantly different at the three temperatures {x± SD=6.8 ±5.8 cm/s at 10°C; 5.3 ±4.8 cm/s at 15°C; and 4.7 ±8.2 cm/s at 20°C). Because ground speed re- mained nearly constant, actual swimming speed in- creased linearly with increasing current speed in the flume (Fig. 2). The distribution of individual swimming velocities within the school was unimodal but was skewed to- wards a speed slightly faster than the flow rate in the flume (Fig. 3 1. The coefficient of variation in swimming speed declined curvilinearly as velocity increased, indicating that fish behavior grew less variable at higher speeds (Fig. 4). The pattern of decreasing variability with increasing swimming speed was similar at all three temperatures. Respiration rates increased exponentially with increasing swimming speed (Fig. 5, Table 1). Analy- sis of covariance revealed that the relationships be- tween respiration rate and swimming speed differed significantly at the three temperatures. The regres- sion slope, or the rate of increase in oxygen consump- tion per unit swimming speed, was greatest at 10 'C, significantly higher than at 15' or 20°C (Fig. 6). The 15° and 20 C curves differed in elevation but not in slope, indicating that the overall level of metabolism was higher at 20°C than at 15 C, but the rate of in- crease in metabolic rate with increasing swimming speed was similar at the two temperatures. The cost of swimming was therefore higher at 10°C (metabo- lism increased by a factor of 2.32 per 1 BL increment in swimming speed) than at 15° or 20°C (metabo- lism increased by 1.75 and 1.65, respectively) (Table 2 ). At the higher test speeds, metabolic rates at 10°C were equal to or greater than those at 15°and 20° (Fig. 6). Standard metabolism, as indicated by they-inter- cepts of the equations given in Table 1, increased with temperature. At 10°, 15°, and 20°C, standard me- tabolism was equal to 0.040, 0.073, and 0.087 mg 0./(g wet wt • h) respectively, or 0.131, 0.238, and 0.284 mg 02/(g dry wt • h). The Qjg for standard me- tabolism (Prosser, 1973) was higher over the inter- val 10-15°C (3.3) than for 15-20°C ( 1.4). The Qj^ over the interval 10-20°C was intermediate (2.2) (Table 3). Discussion Previous work has shown that menhaden swimming behavior reflects environmental conditions and food availability. At 20°C, adult menhaden of about 300 g swim at a characteristic speed of about 12.2 cm/s (0.5 BL/s) in still water in the absence of food (Durbin et al., 1981). During feeding, swimming speed and metabolic rate increase hyperbolically with increas- ing food concentration (Durbin et al., 1981). Hettler ( 1976) found that routine swimming speed in larval and juvenile menhaden (up to 80 g) increased with temperature but decreased with increasing salinity, starvation, and in the dark. Ground speeds in the present experiments were remarkably stable considering the tested tempera- ture range and the different energy expenditures at the different current speeds. The decline in individual variability in swimming speed as mean swimming speed increased (Fig. 4) was very similar to that ob- served by Durbin et al. (1981). This decline in vari- ability may reflect decreased excitability as fish swim faster as well as a transition to steadier swimming with lower energy costs (reviewed in Brett and Groves, 1979; Boisclair and Tang, 1993). The nearly constant ground speed in all trials in- dicated that menhaden did not approach their maxi- mum swimming capacity at any of the tested speeds and, as a result, were probably not fatigued at the end of a day's trials despite the lack of rest between trials. The maximum observed metabolic rate offish in our study, 0.356 mgOy(gwet wt • h) at 20°C, while fish were swimming at 63.2 cm/s, was considerably lower than that associated with filter feeding at a slower speed (0.538 mg O.^ /(g wet wt • h) at 20°C and 43.4 cm/s: Durbin et al., 1981 ), further indicating that the fish were not performing near their physiologi- Macy et al : Metabolic rate of Brevoortia tyrannus 287 70 60 50 40 30 20 10. 70. u fO a- F ct so C/3 ««« o 40 ^ >^ 30 (> a 3 20 rr u r^ 10 0 60 50-1 40 30 20-1 10 0 10°C 13.0 cm/s ttx. 1 r 34.4 cm/s ft-n r 59.8 cm/s a ^ 0 25 50 i. 75 100 15°C 13.8 an/s k. 1 r 35.4 cm/s 1^ 60.8 cm/s 25 r^ T 50 t- 75 — r 100 20°C ■ 12.2 cm/s 2 qi_ 1 1 r 39.5 cm/s In. T r 63.2 cm/s -f^ -r 25 50 75 100 t- Swimming speed, cm/s Figure 3 Frequency distribution of swimming speeds within a menhaden school swimming against typical slow, intermediate, and fast flows, at 10", 15". and 20°C. The flow rate inside the flume is indicated m each graph. Open bars indicate fish swimming against the current faster than the flume speed; diagonally hatched bars indicate fish swimming against the current, but slower than the flume; solid bars indicate fish actively swimming down current (i.e. with the current). cal limit in the present experiments. Hettler (1976) measured a high oxygen consumption rate of 0.82 mg O.ytg wet wt • h) during feeding by juvenile men- haden at 27°C, but the full metabolic scope (Fry, 1957) of menhaden remains unknown. Metabolic scope is related to the surface area of the gills and to the ca- pacity of fish to take up oxygen from the surround- ing medium. The unusually large gill area in men- haden (Gray, 1954) suggests that metabolic scope is also large. We were unable to determine either maximum or critical swimming speeds for adult menhaden (Brett and Glass, 1973) because sufficiently high currents could not be produced in the flume respirometer Current speeds ranged from 12 to 63 cm/s (0.5 to 2.5 BL/s), which spans the reported range of voluntary 288 Fishery Bulletin 97(2), 1999 swimming speeds during routine activity and feeding (Durbin and Durbin, 1975; Hettler 1976; Durbin et al., 1981). Swimming capac- ity is unusually high in Atlantic menhaden; Hartwell and Otto (1978) reported that 5.8- cm-SL juvenile menhaden sustained a critical swimming speed of 15.8 BL/s for 64 min at 20°C. The critical swimming speed of similarly sized sockeye salmon at 15°C is about 7 BIVs (Brett and Glass, 1973). Standard metabolism (Fry, 1957) is a mea- sure of basic maintenance costs and as such may be a useful reference level for interspe- cific comparisons. The magnitude of standard metabolism also seems to reflect species life styles with respect to their general activity levels. Standard metabolic rates for menha- den (Table 3) were lower than those for active predators like the bluefish, Pomatomus saltatrix (0.156 mg 0._,/(g wet wt • h), for 217-g fish at 15 C (Freadman, 1979), but higher than those for the sedentary flatfish Limanda limanda (0.015, 0.029, and 0.042 mg OJ(g wet wt • h) at 5°, 10=, and 15 C for 236-400 g fish (Duthie, 1982)). Standard metabolism in menhaden was simi- lar to that of sockeye salmon (Brett, 1964) at 15^^C, but lower at 10" and 20 = C (Table 3). Durbin et al, ( 1981) previously estimated standard metabolism for menhaden at 20''C to be only 0.036 mg O,^ Ag wet wt ■ h), from the regression of feeding metabolism as a function of swimming speed. However we believe that the present higher estimates are more reliable be- cause the data include a broader range of swimming speeds and require less extrapolation to zero swim- ming speed than the earlier estimate. The QjQ values (Prosser, 1973) provide a measure of the effect of environmental temperature on meta- bolic rates. The Qj,, for standard metabolic rate in our study was very close to that of Hettler for rou- tine metabolism in juvenile menhaden (Table 3). The decline in Q,„ between 10°-15°C and 15''-20°C thus appears to be real. A similar pattern was obsei'X'ed in dab, Limanda limanda, where the Q,,, over 5°- 10°C was higher than over 10 -15°C (3.7 and 2.1 respectively ( Duthie, 1982 ). The more extensive data of Hettler ( 1976) and Brett (1964) indicate that over a wide temperature range (Hettler: 14-24°C; Brett: 5-24°C), the Q,,, for metabolic rate is close to 2.0, but within this range the Q,,j declines as the ther- mal optimum is approached and increases towards the thermal extremes for the species. Hettler's data on routine metabolic rate in juve- nile menhaden, when extrapolated to 300-g fish (Table 3 ), yield relatively high routine metabolic rates compared with those observed by Durbin et al. ( 1981 ) 40 _ , • 10°C A\ O 15°C 30 q\ A 20°C \il c eg 1 20 x-N\ > U e> -^a\^ - 10 1,1,1,1.1.1.1 0 10 20 30 40 50 60 70 Mean SS, cm/s Figure 4 The coefficient of variation (CVl in relation to mean swimming speed (SS) of a menhaden school in the respirometer at 10°, 15°, and 20 C, Table 2 Comparison of the metaboHc cost o f increasi ng swimming speed by 1 BL/sec (ratio of respirat ion rate at 2BL:1 BL/s) | in Atlantic menhaden, Brevoortia t\Tannus (Table 1, this study), with that in sockeye salmon, Oncorhynchus nerka (Table II in Brett 1964) . Fish were not fed for 24 h VIenhaden (this study) Sockeye (Brett, 1964) Temp. Length Wt Length Wt (C) (cm) tg> Cost (cm) (g) Cost 5 18.0 50 2.19 10 26.0 327 2.32 18.0 50 1.91 1.5 25.5 303 1.75 18.0 50 1.86 20 25.5 283 1.65 18.0 50 1.66 24 18.0 50 1.48 (0.10 mg O.J\ g wet wt • h ) at 20°C ). Because the maxi- mum weight of Hettler's fish was only 74.3 g, the extrapolated values for 300-g fish are most likely overestimates. The calculated routine rates in Table 3 correspond to swimming speeds of 18.8, 25.7, and 27.4 cm/s at 10°, 15°, and 20°C, respectively which were faster than the routine speed of 12.2 cm/s at 20' C reported by Durbin et al. (1981 ). Routine metabolic rates are measured directly at spontaneous activity levels. Thus routine rates re- flect real world activity levels, because few fish ac- tually sleep or totally cease movement that would depress metabolic rates to basal or standard levels. Our present observations do seem to be comparable Macy et a!,; Metabolic rate of Brevoortia tyrannus 289 1.000 0.100 0.010 10°C R = 0.040 X e '^ ^^■^ " ^\ r-^ = 0.994 0 10 20 30 40 50 60 70 — 1.000 00 M 0.100 E o. D d* 0.010 15°C R = 0.073 xe'°°22''S)^ ^2^0 898 I I I I , I I I I I I I I I ,, I 0 10 20 30 40 50 60 70 1.000 r 20°C 0.100 R = 0087 xe '0020x5)^^2^0 963 -+n -1^ -1^ r4^ r-l- -A 0.010 0 10 20 30 40 50 60 70 Mean SS, cm/s Figure 5 Relation between metabolic rate and swimming speed offish in the respirometer at 10", 15", and 20°C. to routine rates calculated from Hettler's data (Table 3) that indicate that a 300-g menhaden at 10"'C would have a routine metabolic rate of 0.073 mg 0.,/(g wet wt • h) and to previously cited observations by Durbin et al. ( 1981) on routine swimming and metabolic rates in menhaden at 20°C. Johnstone et al. (1993) re- ported routine metabolic rates of 0.118 mgOV(gwet wt • h) for 290-380 mm Atlantic mackerel. Scomber scombriis, swimming at 0.6 BL/s at 11.1°C, and 0.093 mg 0,/(g wet wt • h) for 255-310 mm Atlantic her- ring, Cliipea harengus, swimming at 0.3 BL/s at 9.3°C. Because these routine swimming speeds were lower than previously reported by these investiga- tors (0.9-1.2 BL/s for mackerel, and 2 BL/s for her- ring) and because the fish were maintained for a pro- longed period in the respirometer without feeding 1.00 [ /^ o ration, mg p o ^ ..■)i^-'* .'■ ■ This Study: IOC -....••;x^" ^,.-'' • 15C jy ,■*' ^ 20C n Brett, 1964: IOC O 15C A 20C 0.02 0 10 20 30 40 50 60 70 80 Swimming speed, cm/s Figure 6 Comparison of respiration rate and swimming-speed rela- tionships in menhaden ( 25.5-26 cm FL. 283-327 g wet wt) (Table 1, our study) and sockeye salmon, Oncorhynchus nerka (IScmTL, 50 g wet wt) (Brett. 1964). (5-16 d for mackerel, 13 d for herring), these may be under-estimates. Routine activity and metabolic rates in menhaden thus may be lower than in her- ring and mackerel. Durbin et al. ( 1981 ) noted that the actual measured routine metabolic rate in menhaden was higher than that predicted for the same swimming speed by us- ing the swimming-speed and metabolic-rate relation- ship during feeding. This higher routine rate was as- sumed to reflect increased excitability in fish when not occupied by feeding and also the more variable and energetically less efficient mode of swimming associated with routine activity. Routine swimming also appears to be energetically more expensive than forced swimming (as shown in our study) and directed swimming, where the optomotor response is used to induce fish to swim at a steady, but possibly more variable, speed than in forced swimming. In a re- view of ten species, three marine and seven fresh- water, Boisclair and Tang ( 1993) found that routine swimming was energetically the most expensive and, on average, was 9.4 times higher (range: 6.4 to 14.0) than forced swimming at the same speed. Directed swimming ranged from 1.0 to 2.8 times as expen- sive. Applying a similar analysis to menhaden at 290 Fishery Bulletin 97(2), 1999 Table 3 Comparison of Qui for standard (Rstj) and routine (R^^^,) metabolism in Atlantic menhaden, Bret'oor/ia tyranniis (this study and Hettler, 1976. respectively), with standard metabolism in sockeye salmon, Oncorhynchus nerka (Table II in Brett, 19641. Hettler's data were extrapolated to a 300-g menhaden by using regressions in his Table I. Temp °C Menhaden (this study) Menhaden (Hettler, 19761 Sockeye Brett, 1964) Temp °C Menhaden (this study) Menhaden (Hettler, 1976) Sockeye (Brett, 1964) Wt R3,, (g) (mg OJ(z ■ h)) Wt (g) '^Rout (mgO/g-h)) Wt °C ^Std (mg O./g ■ h)) 5 50 0.041 10 327 0.040 300 0.073 50 0.060 5-10 2.2 15 303 0.073 300 0.127 50 0.071 10-15 3.3 3.0 1.4 20 283 0.087 300 0.149 50 0.120 15-20 1.4 1.4 2.9 24 50 0.195 20-24 3.4 25 300 0.231 20-25 5-24 10-20 10-25 2.2 2.4 2.0 2.1 2.2 20°C, we first calculated from Table 1 the metabolic rate for forced swimming at the routine speed ( 12.2 cm/s) = 0.111 mgOg/ig wet wt-h) and subtracted the standard metabolic rate from Table 3 (0.087 mg O.J (g wet w • h) to obtain the net cost of swimming = 0.024 mg 0./(g wet wt • h). The net cost of routine swim- ming (Durbin et al, 1981) is 0.1 - 0.087 = 0.013 mg 0.^(g wet wt • h). The ratio of the two (= swimming co"st ratio [SCR] of Boisclair andTang, 1993) = 0.024/ 0.013 = 1.8. This value is much lower than Boisclair and Tang's (1993) average SCR value of 9.4 for the ratio of routine to forced swimming costs and is more comparable to their average SCR value of 1.6 for the ratio of directed to forced swimming costs. This sug- gests that routine activity in menhaden, although energetically more expensive than sustained swim- ming, is relatively economical. Metabolism associated with routine activity is, in fact, only slightly elevated above standard metabolism in menhaden. Menha- den thus may be an exception to the rule (Webb, 1991 ) which states that routine metabolic rates of swim- ming menhaden are not quasi-steady but are sub- stantially higher than forced-swimming rates. The cost of swimming in menhaden increases ex- ponentially with a linear increase in swimming speed (Fig. 5, this study; Durbin et al., 1981), a pattern characteristic offish in general (reviewed by Brett and Groves, 1979). The relation between the meta- bolic cost of swimming and temperature is similar for unfed Atlantic menhaden and sockeye salmon (Oncorhynchus nerka) (Fig. 6; Table 2). In both spe- cies, the rate of increase in respiration rate per in- crement of swimming speed is highest at the lowest temperature. With increasing temperature the slopes of the relationships decrease, indicating that the metabolic cost of swimming declines. The cost of swimming to unfed Atlantic menhaden is close to Beamish's ( 1978 ) mean value for eight species, where metabolism increased by 2.3-fold for a 1 BL/s increase in swimming speed. Thus we conclude that the meta- bolic cost of swimming in nonfeeding menhaden is similar to that of other pelagic fish species. Conversely, the cost of swimming to menhaden while filter-feeding is much higher, where the meta- bolic rate rose five-fold per BL/s increase in swim- ming speed (Eq. 2 in Table 2, Durbin et al., 1981), compared with a 1.65-fold increase in unfed fish at 20 = C (Table 2, our study). Durbin et al. ( 1981) specu- lated that the very high cost of swimming during fil- ter feeding was related to the increased hydrody- namic drag of the expanded gill opercula and the long gill rakers that strain plankton from the water. Present results permit the separation of metabolic rate during feeding into its component parts at 20°C (Table 4; Fig. 7). We observed that during filter feed- ing, menhaden swam at a preferred speed of about 41.3 cm/s over a wide range of plankton concentra- tions. However, after prey concentrations had been reduced to subthreshold levels, feeding ceased and activity and respiration both returned to routine lev- els where there was no postfeeding respiration peak ( Durbin etal., 1981). Standard metabolism accounted for only 18.1'^ of the total metabolism, and the en- ergy demand for swimming was about 22.9%. The remaining 599^ of metabolism was associated with the energy cost of feeding, which includes the in- Macy et al.; Metabolic rate of Brevoortia tyrannus 291 Metabolic budget R No n feed ^Std D "Swim creased drag of the filtration ap- paratus, and the metabohc cost of processing the food (specific dynamic action [SDA]). SDA could not be distinguished from total metabolism during feeding (Durbinetal., 1981) because food is processed very rapidly as was shown by the fact that 50% of ni- trogen was excreted 1.4 h after ingestion (Durbin and Durbin. 1981) and by the lack of a post- feeding respiration peak. The high cost of filter feeding in the menha- den provides an interesting con- trast with a sedentary, ambush predator such as the Northern pike, where the energy required for prey capture is relatively small and where most of the cost of feeding is associated with SDA during an extended postfeeding period of digestion (Lucas et al.. 1991). The bioenergetics of feeding differ between filter feeders, such as menhaden, and particulate feeders that consume zooplankton by individual capture. Prey consumption increases linearly with prey con- centration and predator swimming speed for filter feeders but increases hyperbolically (Durbin, 1979) for particulate feeders because of the limitations of handling time. The energy cost is higher for filter feeding than for particulate feeding. Both feeding modes are energetically more expensive than rou- tine swimming (James and Probyn, 1989). The opti- mal swimming patterns in relation to food concen- tration and foraging time therefore differ between filter and particulate feeders (Ware, 1975; Durbin and Durbin, 1983). Many planktivorous fishes switch between filter and particulate feeding, according to the size and concentration of prey in the water (north- ern anchovy: Leong and O'Connell, 1969; O'Connell, 1972; Hunter and Dorr, 1982; cape anchovy: James and Probyn, 1989; pilchard: van der Lingen, 1994; Pacific mackerel: O'Connell and Zweifel, 1972; At- lantic herring: Gibson and Ezzi, 1985, 1990, 1992). The switch between filter and particulate feeding can be predicted from the relationship between energy gain and expenditure as the plankton size spectrum changes (Crowder, 1985; James et al., 1989; Gibson and Ezzi, 1992). In contrast, Atlantic menhaden are less flexible in their feeding behavior and conse- quently may be restricted to coastal and estuarine waters of high production because energetic costs of filter feeding are high and because a relatively small volume of water is searched with mode of feeding (Durbin and Durbin, 1983). Table 4 Partitioning of metabolism during filter feeding in Atlantic menhaden Brevoortia tyrannus, swimming at their preferred speed of 41.3 cra/s at 20°C. Components of the energy budget: total metabolism during feeding (/Jp^^j); total metabolism nonfeeding but swimming at the same speed '.^Nonfeed'' standard metabolism (^s,j>; metabolism associated with swimming li?s„.,n,>; metabolism associated with fdter feeding l^p,!,^^ '; metabolism associated with digestion and transformation of the food '^sda'- -^Feed fro"i Durbin et al. (1981); remaining data from this study. mg 02/(g wet wt h) Percent of total •'^Std + ^Sw,m + ^Filter + ^SDA ■^Std + ^S»>m = R. 0.480 0.197 0.087 0.110 0.283 100 18.1 22.9 59.0 1000 500 3 00 E 100 10 I ■ 20 ■ I 30 cm/s 40 50 60 ■ I ■ 70 Filtering +SDA 59.0% Swimming speed, BL/s Figure 7 The energy budget of Atlantic menhaden during foraging at the preferred swimming speed. 41.3 cm/s (Durbin et al., 1981) partitioned by the vertical dotted line into the en- ergy required for standard metabolism (horizontal dotted line), swimming (solid line), and feeding (dashed line) = filtering + SDA. However, menhaden flourish in their chosen habi- tat, and as omnivores specializing in the smaller size spectrum of particulate material, they have no im- portant competitors among the fishes. Herring and mackerel eat larger zooplankton and live further off- shore. The small anchovies that live in the men- 292 Fishery Bulletin 97(2), 1999 haden's range are exclusively zooplanktivores (Bige- low and Schroeder, 1953). Mesozooplankton are com- petitors for both food and prey of the menhaden (Peck, 1893; Durbin and Durbin, 1975). Menhaden are im- portant in nearshore food webs as primary herbivores and are a major prey species for predatory fishes ( Bigelow and Schroeder, 1953; Friedland et al., 1989; Baird and Ulanowicz, 1989); because of their great abundance, they support one of the largest domestic fmfisheries in the United States (Ahrenholz et al., 1987). Acknowledgments This research was stimulated by an opportunity granted to two of us (A. and E. Durbin) many years ago to work in the laboratories of William Hettler and David Peters of the National Marine Fisheries Service Laboratory at Beaufort, NC, who were pio- neering the experimental study of menhaden in the laboratory We thank Jonathan Hopkins for the de- sign and construction of the flume respirometer and for his assistance during the experiments. This re- search was supported by NSF grants OCG 89-15610 and OCG91-01985. This paper is dedicated to Ann Durbin who passed away in July 1995. Literature cited Ahrenholz, D. W., W. R. Nelson, and S. P. Epperly. 1987. Population and fushery characteri.stics of Atlantic menhaden. Breioortia tyrannus. Fi.sh. Bull. 85(31:569- 600. Baird, D., and R. E. Ulanowicz. 1989. The seasonal dynamics of the Chesapeake Bay ecosystem. Ecol. Monogr. 59(4):329-364. Barton, B. A., and C. B. Schreck. 1987. Mctahohc cost of acute physical stress in juvenile steelhead. Trans. Am. Fish. ,Soc. 116:257-263. Beamish, F. W. H. 1978. Swimming capacity In W. .S. Hoar and D. J. Randall (eds,). Fish physiology, vol. VII., p. 101-187. Academic Press, New York, NY. Bigelow, H. B., and W. C. Schroeder. 1953. Fishes of the Gulf of Maine. Fish. Bull. 53(74):577. Boisclair, D., and M. Tang. 1993. Empirical analysis of the influence of swimming pat- tern on the net energetic cost of swimming in fishes. J. Fish Biol. 42:169-183. Brett, J. R. 1964. The respiratory metaholism and swimming perfor- mance of young sockeye salmon. J. Fish. Res. Board Can. 21(51:1183-1226. Brett, J. R., and N. R. Glass. 1973. Metabolic rates and critical .swimming speeds of sock- eye salmon iOncorhynchus nerka) in relation to size and temperature. J. Fish. Res. Board Can. 30(3):379-387. Brett, J. R., and T. D. D. Groves. 1979. Chapter 6, Physiological energetics. In W. S. Hoar, D. J. Randall, and J. R. Brett (eds. I, Bionergetics and growth, p. 279-352. Academic Press, New York, NY. Carritt, D. E., and J. H. Carpenter. 1966. Comparison and evaluation of currently employed modifications of the Winkler method for determining dis- solved oxygen in seawater; a NASCO report. J. Mar. Res. 24:286-318. Crowder, L. B. 1985. Optimal foraging and feeding mode shifts in fish. Environ. Biol. Fishes 12(l):57-62. Durbin, A. G. 1979. Food selection by plankton feeding fishes. In H. Clepper (ed.), Predator-prey systems in fisheries manage- ment, p. 203-218. Sport Fishing Institute, Washington, DC. Durbin, A. G., and E. G. Durbin. 1975. Grazing rates of the Atlantic menhaden Brevoortia tyrannus as a function of particle size and concentration. Mar. Biol. 33:26.5-277. 1981. Assimilation efficiency and nitrogen excretion of a filter-feeding planktivore, the Atlantic menhaden, Brevo- ortia tyrannus iPiSces: Clupeidae). Fish. Bull. 79(41:601- 616. 1983. Energy and nitrogen budgets for the Atlantic men- haden, Brevoortia tyrannus (Pisces: Clupeidael. a filter- feeding planktivore' Fish. Bull. 81(21:177-199. Durbin, A. G., E. G. Durbin, P. G. Verity, and T. J. Smayda. 1981. Voluntary swimming speeds and respiration rates of a filter-feeding planktivore, the Atlantic menhaden, Brevoortia tyrannus (Pisces: Clupeidae). Fish. Bull. 78(4):877-886. Duthie, G. G. 1982. The respiratory metabolism of temperature-adapted flatfish at rest and during swimming activity and the use of anaerobic metabolism at moderate swimming speeds. J. Exp. Biol. 97:359-373. Frcadman, M. A. 1979. Swimming energetics of striped bass iMorone saxatilis) and bluefish ( Pomatomus saltatnx I: gill ventilation and swim- ming metabolism. J. Exp. Biol. 83:217-230. Friedland, K. D., L. W. Haas, and J. V. Merriner. 1984. Filtering rates of the juvenile Atlantic menhaden Brevoortia tyrannus iPiscefi: Clupeidae), with consideration of the effects of detritus and swimming speed. Mar Biol. 84:109-117, Friedland, K. D., D. W. Ahrenholz, and J. F. Guthrie. 1989. Influence of plankton on distribution patterns of the filter-feeder Breroo;7!a tyrannus (Pisces: Clupeidael. Mar. Ecol. Prog. Ser 54(1&2):1-11. Fry. F. E. J. 1957. The aquatic respiration offish, /n M. E. Brown (ed.). The physiology of fishes, vol. 1. Academic Press, New York, p. 1-79. Gibson, R. N., and I. A. Ezzi. 1985. Effect of particle concentration on filter- and particu- latc-feeding in the herring Clupea liarengus. Mar. Biol. 88:109-116. 1990. Relative importance of prey size and concentration in determining the feeding behaviour of the herring Clupea harengus. Mar. Biol. 107:357-362. 1992. The relative profitability of particulate- and filter- feeding in the herring, Clupea harengus L. J. Fish Biol. 4O(4):577-590. Macy et aL: Metabolic rate of Brevoortia tyrannus 293 Gray, I. E. 1954. Comparative study of the gill area of marine fishes. Biol. Bull. 107:219-226. Hartwell, S. I., and R. G. Otto. 1978. Swimming performance of juvenile menhaden (Brevoortia tyrannus). Trans. Am. Fish. Soc. 107(61:793-798. Hettler, W. F. 1976. Influence of temperature and salinity on routine metabolic rate and growth of young Atlantic menhaden. J. Fish Biol. 8:55-65. Hunter, J. R., and H. Dorr. 1982. Thresholds for filter feeding in northern anchovy, Engraulis mordax. CalCOFI Rep. 23:198-204. James, A. G., and T. Probyn. 1989. The relationship between respiration rate, swimming speed and feeding behaviour in the Cape anchovy, Engraulis capensis Gilchrist. J. Exp. Mar. Biol. Ecol. 131:81-100. James, A. G., T. Probyn, and L. Hutchings. 1989. Laboratory-derived carbon and nitrogen budgets for the omnivorous planktivore Engr'aulis capensis Gilchrist. J. Exp. Mar Biol. Ecol. 131:125-145. Johnstone, A. D. F., C. S. Wardle, and S. M. Almatar. 1993. Routine respiration rates of Atlantic mackerel, Scomber scombrus L., and herring, Clupea harengus L., at low activity levels. J. Fish Biol. 42:149-151. June, F. C., and F. T. Carlson. 1971. Food of young Atlantic menhaden, Brevoortia tyrannus. in relation to metamorphosis. Fish. Bull. 68(3):493-512. Kanda, T., and Y. Itazawa. 1978. Group effects on physiological and ecological phenom- ena in fish-II group effect on the growth of medaka. Bull. Jpn. Soc. Sci. Fish. 44(11):1197-1200 (Engl, abstract.) Leong, R. J. H., and C. P. O'Connell. 1969. A laboratory study of particulate and filter feeding of the northern anchovy (Engraulis mordax I. J. Fish. Res. Board Can. 26(3):557-582. Lucas, M. C, L G. Priede, J. D. Armstrong, A. N. Z. Gindy, and L. De Vera. 1991. Direct measurements of metabolism, activity and feeding behaviour of pike, Esox lucius L., in the wild, by the use of heart rate telemetry. J. Fish. Biol. 39:325-345. O'Connell, C. P. 1972. The interrelation of biting and filtering in the feed- ing activity of the northern anchovy (Engraulis mordax). J. Fish. Res. Board Can. 29(3):285-293. O'Connell, C. P., and J. R. Zweifel. 1972. A laboratory study of particulate and filter feeding of the Pacific mackerel. Scomber japonicus. Fish. Bull. 70(3):973-981. Peck, J. I. 1893. On the food of the menhaden. Bull. U.S. Fish. Comm. 13:113-126. Prosser, C. L. 1973. Chapter 9: Temperature, /ri C. L. Prosser, (ed.), Com- parative animal physiology, p. 362-428. W.B. Saunders, Philadelphia, PA. Ross, R. M. and T. W. H. Backman. 1992. Group-size-mediated metabolic rate reduction in the American shad. Trans. Am. Fish. Soc. 121:385-390. Sanderson, S. L., and J. J. Cech Jr. 1992. Energetic cost of suspension feeding versus particu- late feeding by juvenile Sacramento blackfish. Trans. Am. Fish. Soc. 121:149-157. Skazhina, E. P. 1975. Energy metabolism of anchovy Engraulis encra- sicholus L. when kept in a group or isolated and when anes- thetized with quinaldine. Doklady Biological Sciences. Engl. Transl. (1976). Doklady Akademii Nauk SSSR. 225il):238-240. Strickland, J. D. H., and T. R. Parsons. 1972. A practical handbook of seawater analysis. Bulletin, Fisheries Research Board of Canada, 167, Fisheries Re- search Board of Canada, Ottawa, 310 p. van der Lingen, C. D. 1994. Effect of size and concentration on the feeding behaviour of adult pilchard Sardinops sagax. Mar. Ecol. Prog. Ser 109(1):1-13. Ware, D. M. 1975. Growth, metabolism, and optimal swimming speed of a pelagic fish. J. Fish. Res. Board Can. 32(1);33-41. Waring, C. P., R. M. Stagg, and M. G. Poxton. 1992. The effects of handling on {\ounder (Platichthys flesus L.I and Atlantic salmon (Salmo salar L.). J. Fish Biol. 41:131-144. Webb, P. W. 1991. Composition and mechanics of routine swimming of rainbow trout, Oncorhynchus mykiss. Can. J. Fish. Aquat. Sci. 48(4):583-590. 294 Abstract.— We examined the relation- ship between otolith size and larval standard length (SLl at hatching for Atlantic cod. Gadus morhua, on the Scotian Shelf. We found a weak corre- lation between SL and area-based and radius-based measures of size of both lapillar and sagittal otoliths. Correla- tions of SL with the area of the lapillus were strongest. However, the predictive ability of all relationships was low. For example, the range of predicted SL at hatching of larvae with lapilli of aver- age area included more than 90'^t of observed SLs in newly hatched larvae from the Scotian Shelf collected over two spawning seasons. These results suggest that otolith-based attempts to backcalculate the size of cod larvae may be prone to substantial error if sizes of particularly young larvae are esti- mated. We recommend that, where pos- sible, stock- and season-specific esti- mates of the relationship between the area of the lapillus and larval size at hatching be used in back-calculation techniques. The relation between otolith size and larval size at hatching for Atlantic cod, Gadus morhua* Thomas J. Miller Chesapeake Biological Laboratory University of Maryland Center for Environmental Science PO Box 38 Solomons, Maryland 20688-0038 E-mail address: millerg cbl umces edu Tomasz Herra William C. Leggett Department of Biology Queens University Kingston, Ontario, K7L 3N6 Canada Manuscript accepted 22 June 1998, Fish. Bull. :294-305( 1999), The variation in year-class strength in fish populations has profound implications for our ability to man- age stocks wisely. Recently, consid- erable effort has been focused on trying to understand why the few individuals that survive do so, rather than why the majority die (Crowder et al., 1992), This ap- proach, which relies inherently on characterizing traits in individual fish, has been widely applied in both marine and freshwater studies (Herman et al,, 1996; Rice et al., 1997 for example). The philosophy behind the approach is that by un- derstanding how survivors differ from those that die, one may expose the mechanisms that regulate re- cruitment (Fritz et al,, 1990). We used this approach to explore recruitment mechanisms in Atlan- tic cod, Gadus morhua, on the Scotian Shelf, In this area, cod spawning is bimodal, beginning in late October and continuing to the following April, with peaks in De- cember and March (Miller et al,, 1995), Recent studies have sug- gested the importance of small-scale physical oceanographic features, such as gyres and fronts, for larval survival (Taggart et al., 1996; Lochmann et al,, 1997). Abundant populations of copepods, particu- larly F^seudocalanus and Paraca- lanus, co-occur with cod larvae ( McLaren and Avendano, 1995 ) and support rapid larval growth. By fol- lowing birth-date cohorts through time and repeatedly estimating the distribution of phenotypic and ge- notypic traits in the cohort, we hoped to quantify whether survi- vors were different functionally from the majority that died, or whether they were simply lucky (Miller, 1997), Genetic evidence suggests that cod larvae collected on the Scotian Shelf originate from distinct spawn- ing events (Ruzzante et al., 1996). However, Ruzzante et al. (1996) have shown that the genetic struc- ture within the population remains stable over time. Meekan and Fortier (1996) repeatedly sampled two autumn-spawned cohorts of Scotian Shelf cod, following each for approximately six months. The pat- tern of survival for the two cohorts differed. In 1991-92, the growth rate distribution of survivors dif- fered little from the growth distri- bution of the cohort from which they were drawn. In contrast, in 1992- 93, Meekan and Fortier (1996) found evidence of strong selection * Contribution .3013 from the LIniversity of Maryland Center for Environmental Sci- ence. Solomons. MD. Ml iller et al : Relation between otolith size and larval size at hatching for Gadus morhua 295 N for faster growing fish. Moreover, after comparing otolith sizes at hatching, Meekan and Fortier ( 1996) suggested that the potential for faster growth ex- pressed by the survivors may have been present at hatching. Our ability to detect examples of phenotypic selec- tion evidenced above depends on our ability to quan- tify the distribution of traits in the entire cohort and to hindcast the distribution of traits in the survivors (Miller, 1997). For much of the research discussed above, and for many other individual-based studies, otolith microstructural analysis is used to detect phenotypic selection. However, to apply this ap- proach, three requirements must be satisfied (Francis, 1990): 1) primary increments must be de- posited at a known and consistent rate; 2) there must be a quantifiable relationship between growth and the width of increment rings; and 3) the initial otolith size, defined by the presence of a check mark, must be re- lated to the size of the fish at the formation of the check. For cod, the regularity of increment deposition has been verified and validated ( Bergstad, 1984; Radtke, 1984; and see Geffen, 1995, for a recent example). Thus, the first condition for back calculation has been satisfied in cod. However, there remains consider- able uncertainty over the status of the remaining two conditions. The relationship between somatic and otolith growth in cod is unclear ( see Geffen, 1995, and Meekan, 1997, for opposing views). The lack of a clear relationship between rates of so- matic and otolith growth may result from the natural variability among and within many cod populations (Brander, 1994; Chambers, 1997). Typcially, cod larvae grow at highly variable rates, and somatic and otolith growth may become disasso- ciated (Suthers et al., 1989; Cam- pana and Hurley, 1989; Geffen, 1995). Concerns over the validity of the assumed relationship between fish and otolith size at hatching also arises because of inter- and intrapo- pulation variation. Considerable variation has been reported in lar- val size at hatching both within and among populations (Bolz and Lough, 1983; Bergstad, 1984; Radtke, 1984; Knutsen and Tilseth, 1985; Miller et al., 1995). Currently, no clear link between otolith size and larval size at hatching has been established for any population. The objective of this paper is to examine the rela- tionship between otolith size at hatching and larval size at hatching for cod. Specifically, we address whether otolith size at hatching is significantly cor- related with larval size at hatching, and whether all otoliths provide an equally accurate and precise es- timate of larval size at hatching. We use data col- lected for cod on the Scotian shelf collected between 1991-93 to address these questions. Methods Sampling was carried out during 29 cruises on the Scotian Shelf between March 1991 and May 1993 (Fig. 1). O'Boyle et al. (1984) have given a general description of the Scotian Shelf system. Full details of the sampling method are given by Miller et al. (1995). We summarize the general sampling design here and provide additional details that relate to the specific objectives of this study. Twenty six cruises were designed to provide broad- scale information on temporal and spatial distribu- tions and abundance of cod eggs, larvae, and juve- niles (Fig. 1). We sampled a rectangular grid of 45 44 - 43 63 62 61 60W Figure 1 Map of Scotian Shelf showing the area of sampling locations. Shown on the figure are the 50- and 100-m depth contours, the principal sampling loca- tions within the 15-station grid (open circles), and the location of Halifax, NS, Canada, for reference. 296 Fishery Bulletin 97(2), 1999 stations at roughly monthly intervals using either paired 0.6 1-m bongo nets fitted with 333-|im mesh nets, an 8 + 2 m rectangular mid-water trawl (RMT) fitted with 1600-|i m and 333-|j m mesh nets, or a paired 1.4 m'^ rectangular frame net, fitted with 333-|am mesh nets. Depth information for both the bongo and frame net was estimated from cable angles and lengths deployed. The RMT was fully equipped and provided continuous, real time depth, temperature, salinity, and volume filtered data. The station exhibiting the highest concentration of cod larvae was resampled with a BIONESS sampler equipped with ten 1-m^, 333-|im mesh nets, that was deployed to sample dis- crete 5-m depth strata in the upper 25 m of the wa- ter column and 10-m depth strata at deeper depths. Three of the 29 cruises were designed to track a patch of eggs and larvae over smaller spatial scales for up to 20 days in order to track how traits changed over time. On these cruises we deployed principally a BIONESS sampler and bongo nets. However, because we were attempting to sample continuously from the same patch of water, stations were distributed irregu- larly in space. All net samples were sorted and cod eggs were re- moved on board ship. Late-stage eggs that appeared healthy and undamaged by the collection process were videotaped under a dissecting microscope at 6— 50x magnification. Individual eggs were incubated separately on a 12-h light: 12-h dark cycle and at near- ambient temperature. Light from blue incandescent bulbs mimicked the light environment at depth. Nursery temperatures were recorded daily. All eggs from the same cruise were incubated at the same temperature. However, because sea temperatures varied across the sampling grid, and with depth, there were unavoidable differences between incuba- tion and ambient temperatures for individual eggs. Vials were checked every 12 hours for hatching. When a larva hatched, it was immediately videotaped and stored in liquid nitrogen prior to otolith extraction and analysis. In the laboratory, lapillar and sagittal otoliths were removed from larvae and mounted in cyanoacrylic cement. In most larvae we could remove and classify successfully all four otoliths. Otoliths were examined under bright field illumination under a compound microscope at 60-1 OOOx magnification. Oil immer- sion was required for the higher magnifications. Videotape recordings and otolith images were ana- lyzed in the laboratory by using an image analysis sys- tem (Optimas v3.11, Bioscan Corporation, Seattle WA). We staged each egg according to Thompson and Riley's (1981) system. Egg diameters were calculated from three digitized points on the circumferences of eggs. From these points the diameter was calculated as Diameter ■ a be 4^s{s - a)(s - b){s - c) where a. b, and c = the lengths of the chords con- necting the three points; and s - 2{a + b + c). Larvae that hatched from these eggs were measured to standard length (SL). Only undamaged otoliths were analyzed. Several measures were taken to describe otolith size. At this early stage, otoliths were essen- tially spherical in cross section. We measured the cross- sectioned area of both lapillar and sagittal otoliths. We also measured the radius of the otolith at hatching be- cause this is the most common measurement used when estimating larval hatching sizes from otolith data. Analysis of the data collected depended upon the purpose to which the analysis was being put. For univariate analyses to determine seasonal trends in a trait, data were aggregated to provide a mean value for each deployment before analysis. This approach reflects the sampling design employed in the field, and thus the deployment is the appropriate sampling unit. However, for bivariate analyses to determine the correlation among measures of otolith size and between otolith size and fish size, the individual fish is the appropriate sampling unit, and thus these analyses were conducted at the individual level. Results We identified and incubated 650 cod eggs from April 1991 to May 1993. Of this total, 259 (39.9%) success- fully hatched. Otoliths from a random sample of 73 of these larvae were used in our analyses. We ob- tained reliable measurements from both lapilli on 56 larvae (76.7%) and from both sagittae for 59 (80.1%) larvae. The distribution of data, by month and year, is given in Table 1. We examined the correlation structure in the data for area of otolith and radius of otolith at hatching. Estimates of otolith area were correlated among otolith types, but there were no significant correla- tions among otoliths from the same side of the body (Fig. 2). Moreover, the correlation among lapilli was greater than that for sagittae (Fig. 2). An identical pattern was found with respect to radius of otolith at hatching (Fig. 3). We regressed each measure of otolith size on lar- val size at hatching, using only otoliths from the left side of the body. The area of the lapillus (LA) and SL at hatching were significantly and positively related (Fig 4A). Overall, longer larvae have bigger lapilli. The 95% CIs around the predicted mean were nar- Miller et al : Relation between otolith size and larval size at hatching for Gadus morhua 297 Table 1 Mean size ( ±SD ) and number of newly hatched cod larvae from the Scotian Shelf prov ding data for otolith analysis by month and year. Otolith mean and SDs are calculated by using estimates for both otoliths of the specified type within each fish. Shown for each entry are the mean (upper number), SD (middle num ber in parentheses), and number of fish providi ng data for analyses 1 (lower number). Measurement 1991 -92 1992 -93 Nov Dec Jan Mar Oct Dec Mar Total Egg diameter (mm) 1.35 1.45 1.35 1.51 1.38 1.51 1.59 1.47 (0.1) (0.07) (0.05) (0.1) (0.06) (0.06) (0.09) (0.1) 43 37 11 2 91 40 35 259 SL at hatching (mm) 4.31 4.43 5.04 5.34 4.03 4.52 5.03 4.38 (0.2) (0.3) (0.2) (0.3) (0.3) (0.3) (0.3) (0.5) 43 37 11 2 91 40 35 259 Lapillar area i\im-) 493.0 529.5 696.49 732 360.69 562.51 584.48 537.34 (113.9) (66.1) (98.5) (74.0) (113.2) (90.5) (131.6) 6 7 5 1 10 8 19 56 Lapillar radius at hatching (|im) 13.6 14.12 15.87 15.87 11.71 14.80 14.74 14.13 (1.4) (0.8) (1.1) (0.9) (1.2) (1.0) (1.6) 6 7 5 1 10 8 19 56 Sagittal area (|im^) 191.3 270.29 369.12 360.57 227.25 333.37 280.71 276.6 (69.6) (83.1) (108.6) (50.9) (84.9) (64.1) (85.9) 6 11 5 1 10 8 18 59 Sagittal radius at hatching (Hm) 8.79 10.39 11.98 11.85 9.55 11.32 10.64 10.45 (1.29) (1.4) (1.7) (1.2) (1.5) (1.3) (1.5) 6 11 5 1 10 8 19 60 row. For a lapillus of average area, the preiiicted mean larval SL at hatching was 4.55 mm. The 951 CIs around this value were 4.4 mm < SL < 4.66 mm, a range of 0.2 mm. However, in our application we were more interested in the 95^7^ CIs of an individual prediction. The range in these values was much wider, 3.66 mm < SL < 5.41 mm., a range of L75 mm. We conducted a similar analysis for the radius of the lapillus at hatching (LR, Fig 4B), which was significantly and positively related to SL at hatch- ing. Overall, longer larvae had wide lapilli at hatch- ing. As with the results for the total area of the lapil- lus, the 95'7f CIs for larval SL predicted for a lapillus of average radius were narrow. For a lapillus of av- erage radius at hatching (14.15 |im), the associated 95'7f CIs of the mean were 4.44 < SL < 4.69 mm, a range of 0.25 mm. However, the prediction is less precise for individual back calculations for the asso- ciated 95*7^ CIs of the individual prediction were 3.72 < SL < 5.49 mm, a range of 1.77 mm. We conducted a similar regression analysis for the area and radius of the sagittae and SL at hatching. There were significant linear relationships for both measures of otolith size and SL at hatching (Fig. 5). For an average sagittal area of 276 |im, the predicted SL at hatching was the same as that for the lapillus. The 95% CIs for the population average were 4.38 mm < SL < 4.75, a range of 0.37 mm. The wider con- fidence intervals for the population mean for the sag- ittal measurements compared with the lapillar mea- surements reflected the lower coefficient of determi- nation ir") of the regression. Moreover, the 95% CIs of an individual prediction, based upon sagittal area, were also wider than those for lapillus-based predic- tions (3.48 < SL < 5.64, a range of 2.16 mm). Similar patterns were observed for the regressions for sagit- tal radius at hatching (Fig. 5B). We examined the potential for egg size and tem- perature to increase the predictive power of the re- lationships. We performed these analyses on data aggregated to the deployment level. Standard length at hatching was positively related to egg size: SL = 0.656 (±0.46) -1- 2.55 (±0.31) x egg diameter, n = 128, r~ = 0.345, P = 0.0001. To explore the potential for egg size to explain additional variation in otolith- size and SL-at-hatching models, we regressed the residuals from the relationship of egg diameter to SL on several measures of otolith size. If egg size explains additional variation, independent of SL at hatching, we would expect to see a significant re- gression statistic. The residuals were significantly related to the area and the radius at hatching of the 298 Fishery Bulletin 97(2), 1999 r=0.53. p<0 01 , n=49 •: V r=0.69, p<:0,001. n=59 • -s r=0.68. p<0.001, n=53 700 R.sagitta . ..f- r=0.86, p < 0.001 , n=56 r=0.60, p<0.001, n=50 ' r=0.66, p<0.0C1, n=54 600 L.sagitta .; .•/• • .; 900 200 R.lapillus J L 100 800 Llapillus (200 I Figure 2 Scatter plots of individual measures of otolith area (|jm-) for newly hatched cod larvae collected on the Scotian Shelf Correlation coefficients (/) for each relationship are given in each panel. lapilli (Fig. 6). However, the residuals were not sig- nificantly related to either measure for sagittae. Standard length (SL) at hatching was negatively related to sea temperature at collection: SL = 5.26 (±0.07) - 0.939(±0.007) X temp, n = 44, r- = 0.79, P = 0.001. To explore the potential relationship between otolith size and temperature, we regressed our most predictive estimate of otolith size, lapillar area, against the residuals from the SL-temperature rela- tionship. There was no clear relation between tem- perature residual and lapillar area (Fig. 7). Discussion Otoliths can be used to infer the standard length of cod larvae at hatching. However, our data suggest that the precision with which such inferences can be drawn may not be sufficient to permit accurate com- parisons of the size distribution at hatching between survivors and the population at large. On the Scotian Shelf newly hatched cod larvae varied between 2.4 and 6.1 mm SL (Miller et al., 1995). The best regres- sion relationship we developed in our study related SL to area of the lapillus, and explained 359f of the variation in the data. Accordingly, 65% of the varia- tion remained unexplained. The regressions we de- veloped suggest that the 95*^ confidence intervals of individual back-calculated size at hatching for a fish are wide. Predicted SL for larvae with average-size lapilli ranged from 3.66 to 5.41 mm. This covers fully 48.7'7c of the total range in initial sizes at hatching observed in cod larvae from the Scotian Shelf More strikingly, based on estimates of mean and variation of SL given by Miller et al. ( 1995), this range includes 917( of all newly hatched cod larvae on the Scotian Shelf Back calculations from lapillar otoliths larger or smaller than average size will be even more im- precise. Thus, we conclude that estimates of initial size in cod larvae are unlikely to be sufficiently pre- Miller et al.: Relation between otolith size and larval size at hatching for Gadus morhua 299 r=0.62, p<0.001, n=49 •J • r=0.62, p<0.001, n=59 •• • r=0.48, p<0.01, n=53 ; J.- 18 R.sagitta ° 1 1 1 1 1 • : r=0.80.p< 0.001. n=56 r=0.44, p<0.05, n=50 18 R.lapillus 6 1 1 1 1 1 r=0.36. p<0.05. ffcS4 18 L.sagitta 6 1 1 1 1 1 18 L.lapillus |6 1 1 1 1 1 Figure 3 Scatter plots of individual measures of otolith diameter (mm) at hatching for newly hatched cod larvae collected on the Scotian Shelf Correlation coefficients (r) and sig- nificance for each relationship are given in each panel. cise to detect anything other than substantial differ- ences in hatching size among the survivors and the initial cohort. Meekan and Fortier ( 1996) reported evidence that surviving cod larvae on the Scotian Shelf in 1992-93 were significantly different from the cohort from which they originated. In contrast, survivors did not differ from the cohort at large in 1991-92. Meekan and Fortier (1996) suggested that the change in ra- dius at hatching that they observed in 1992-93 was evidence of selection for faster growing larvae in that cohort. Meekan and Fortier's suggestion, as it relates to the relative sizes of otoliths in the survivors and the initial cohort, is indisputable. Our findings that the size of a larva and its otoliths are weakly corre- lated do not allow us to reject the notion that survi- vors may have differed in size at hatching from the overall cohort. Our results suggest an additional ex- planation. We report here a significant relationship between otolith size and the residual from the pre- dicted relationship between egg and larval sizes. Larvae that hatch from relatively larger eggs, which have relatively larger yolks, have larger otoliths. Hence, we agree with the Meekan and Fortier's ( 1996) suggestion, but we would further hypothesize that the difference observed may have been caused by variation in egg size rather than by variation in larval size at hatching directly. If correct, we suggest that the greater amount of yolk in larger eggs enhanced initial larval growth and survival. An understanding of the actual mechanism responsible for these faster growing larvae will require additional analyses. Our results suggest that attempts to backcalculate size of cod larvae will be most successful if they are based on measurements from the lapillus. Although larval SL at hatching was significantly related to all measures of otolith size, relationships involving the lapillus explained more variation than correspond- ing relationships involving the sagitta. Moreover, regressions involving the projected area of the lapil- 300 Fishery Bulletin 97(2), 1999 65 6 5.5 5 4.5 4 35 3 E 25 O OqOc ^ 200 0) 300 400 500 600 700 Laplllar area (fjm^) 800 900 3 ■ 2.5 - 10 12 14 16 18 Radius of lapillus at hatching (|im) 20 Figure 4 Relation between standard length (SLi at hatching of individual larvae and (Ai the lapillar area iLAl. and (B) the lapillar radius at hatching (LRi for newly hatched cod larvae on the Scotian Shelf Shown on each plot are the predicted linear relationships and 95'; confidence intervals of the mean and of individual predictions. Regression relationships are LA = 3.295 (+0.24) + 0.0023 (±0.0004) x SL, n = 54, r- = 0.35, P = 0.0001; LR = 1.898 l± 0.51) + 0.188 1+0.036) V SL. n = 54, r- = 0.35, P = 0.0001. lus were more accurate than those involving the ra- dius of the lapillus at hatching. We conclude from these findings that attempts to hindcast size at hatch- ing in cod should rely on measurements taken from the lapillus and, where possible, should use the cross- sectional area of the otolith as the measure of otolith size. It has been recognized thpt cod lapilli are larger initially than sagittae, but that sagittae increase in size more rapidly (Bergstad, 1984; Radtke, 1984; Campana and Hurley, 1989). Lapilli remain larger than sagittae for up to 25 days. As a result, back cal- culations involving cod older than 25 d or larger than 6-8 mm have been based on sagittae instead of lapilli. However, the relative sizes of the otoliths at their core remain unaltered by subsequent growth dynam- ics. Hence, we caution that, even though the sagittae in these larger and older fish are larger absolutely than the lapilli, the precision of estimates of initial size at hatching will be most precise if the estimates are based on the size of the lapillus at hatching. In controlled laboratory experiments, Geffen ( 1995) found that the otolith-size and fish-size relationship Miller et al.: Relation between otolith size and larval size at hatching for Gadus morhua 301 100 200 300 400 Sagittal area (nm^) 500 600 65 r 6 55 5 45 4 35 ■ 3 25 B O °o o °% '-ooj^'--- -r?"- 10 12 14 Radius of sagitta at hatching (urn) 16 Figure S Relation between size at hatching of individual larvae and (A) the sag- ittal area (SA), and (B) the sagittal radius (SR) at hatching for newly hatched cod larvae on the Scotian Shelf. Shown on each plot are the predicted linear relationships and 95% confidence intervals of the mean and of individual predictions. Regression relationships are SA = 4.033 (±0.21) + 0.00178 (±0.0007) X SL, n = 58, r^ = 0.09, P = 0.017; SR = 3.44 (±0.45) + 0.104 (±0.043) x SL, n = 59, r- = 0.09, P = 0.018. could predict overall mean population growth rates extremely well but that they were unreliable predic- tors of individual larval growth. For example, Geffen (1995) reported that within 15 d of hatching, otolith growth and a measure of somatic growth were poorly correlated in cod larvae. If Geffen's conclusion reflects a broad pattern, her results have profound implica- tions for any application of an individual-based back- calculation approach to cod. Recently, Meekan ( 1997) pointed out that Geffen's conclusions may have been affected by experimental conditions and by the use of a single mean value of larval size at hatching as the origin of the relationship between larval size and otolith size. Larval size at hatching varies substantially in cod (Knutsen and Tilseth, 1985; Miller et al., 1995), and thus Geffen's use of a single size at hatching is unrealistic and cannot be supported by the data avail- able. Yet it is not clear what value one should use for the origin in backcalculating growth trajectories. Our data suggest that the wide variability in the relationship between otolith size and hatching size observed in cod will likely plague attempts to 302 Fishery Bulletin 97(2), 1999 <3> C 0 C a 200 300 400 500 600 700 800 Lapillar area (^lm^) 900 10 12 14 16 1£ Radius of lapillus at hatching (nm) Figure 6 Relation between the residuals from the mean SL and mean egg diam- eter and (A) the lapillar area (LAi and (B) the lapillar radius at hatch- ing (LRi for newly hatched cod larvae on the Scotian Shelf. Shown on each plot are the predicted linear relationships and 95'"! confidence intervals of the mean and of individual predictions. Regression rela- tionships are; Residual = 0.00162 (±0.0003) x LA - 0.721 (±0.186), ;; = 37, ;•- = 0.38. P = 0.0001; Residua! = 0.126 (±0.03) x Lfl - 1.633 (±0.39), n = 37. r- = 0.37, P = 0.0001. backcalculate to early periods of the life history. Geffen (1995) commented on the variability in the sizes of hatchmarks on the lapilli of cod larvae. She reported estimate.s of otolith diameters at hatching that varied by more than twofold ( 13.7-28.4 |im). In Figure 8, we summarize published data on SL and lapillar diameter at hatching for a range of cod .stocks. The variability evident in Figure 8 suggests that the choice of a "biological intercept" for cod is problem- atical (Campana, 1990); a single "best" size at hatch- ing or otolith size at hatching clearly cannot be de- fined for cod. However, the data do show that an over- all relationship between the mean SL at hatching and the mean diameter of lapillus exists. However, given the limited data set, the relationship is statis- tically insignificant (MDL = 1.73 xMSL -(-19.54, /■'-=0.306, n=7, P>0.05). With these results, we rec- ommend that extreme caution be exercised when selecting parameter estimates for use in the back calculation of size-at-age in cod. Miller et al.: Relation between otolith size and larval size at hatching for Gadus morhua 303 c -.- --^--o-o- - o 10 12 14 16 1£ Radius of laplllus at hatching (|im) 20 Figure 7 Relation between the residuals from mean SL and mean water tem- perature at collection and (A) the area of the lapillus (LA) and (B) the lapillar radius at hatching (LR) for newly hatched cod larvae on the Scotian Shelf Shown on each plot are the predicted linear relation- ships and 95'7f confidence intervals of the mean and of individual pre- dictions. Regression relationships are Residual = 0.00021 (±0.00027) x LA - 0.102 (± 0.143), n = 37, r- = 0.01, P = 0.89; and Residual = 0.013 (±0..21) X i,fl- 0.184 (±0.28), /! =37.;-'- = 0.001, P = 0.85. The implications from our results for studies on other species are less clear It is important to note that one would not expect all species to exhibit the same plasticity in size at hatching, or in the rela- tionship between fish size and otolith size at hatch- ing. Hence, for some species there is little variation in initial size at hatching. In these species the error introduced by assuming a single, universal size at : hatching is probably small. However, it is important to recognize that back-calculation techniques all as- sume a common origin for the family of size-at-age lines that are modeled. The methods differ only in the definition of the origin (fish size at zero otolith size, biological intercept, or otolith size at zero fish size), but all assume a single value. Indeed Campana ( 1990) noted that relatively little attention had been paid to the effects of variation in the intercept term of back-calculation methods; most attention had been paid to variation in the slope. However, for species that are known to vary widely in hatching size (e.g. 304 Fishery Bulletin 97(2), 1999 35 4 45 Standard length at hatching (mm) Figure 8 Relation between larval size and otolith size at hatching reported in the literature for different popula- tions of cod. Shown for each data point are means and bivariate standard deviations (when available). The sources for individual data points are shown as letter codes above the abscissa. Codes: (Al Radtke ( 1984); (Bl Bergstad (1984); (C) Radtke il989l; (D) Geffen (1995); (E) 1991-92 cohort from Meekan and Fortier (19961; (F) 1992-93 cohort from Meekan and Fortier (1996); and (G) this study. Estimates of SL at hatch- ing for Meekan and Fortier were taken from estimates of mean hatching size for each cohort given by Miller et al. (1995). Atlantic herring), individual variability in the inter- cept may be significant. We recommend that, at a minimum, stock specific values for intercepts be used if the interest is in periods of the life history shortly after hatching. Further, caution should be exercised in relating shifts in the back-calculated distribution of radii at hatching to size-selective mortality. More careful study of the relationship between otolith size and fish size in individual species is likely warranted if researchers wish to employ otolith-based back cal- culation (Chambers and Miller, 1995). and the crew of the RV Petrel V for their help. Fund- ing for this research was provided by OPEN (the Ocean Production Enhancement Network), one of 15 Centres of Excellence supported by the Government of Canada, through the Natural Sciences and Engi- neering Research Council (NSERC) of Canada, and by an NSERC Strategic research grant to WCL. TJM was supported in part by the U.S. National Science Foundation under grant no. OCE-9728750. Literature cited Acknowledgments We thank Francois Landry, Rob Douglas, and Patricia Avendano for help in videotaping cod eggs and newly hatched larvae at sea. The cruises that provided these data could not have been conducted without the as- sistance of a team of colleagues from Dalhousie and Laval Universities. We thank Capt. Wayne Walters Bergstad, O. A. 1984. A relationship between the number of growth incre- ments on the otoliths and age of larval and juvenile cod, Gadus morhua L. Flodevigen Rapportser 1:251-272. Brander, K. 1994. Patterns of distribution, spawning, and growth in North Atlantic cod: the utility of inter-regional compari- sons. ICES Mar Sci. Synip. 198:406-413. Bolz, G. R., and R. G. Lough. 198.3. (Irowth of larval Atlantic cod, Gadus morhua. and Miller et al,: Relation between otolith size and larval size at hatching for Gadus morhua 305 haddock, Melanogrammus aeglefmus. on Georges Bank, spring 1981. Fish. Bull. 81;827-8.36. Campana, S. E. 1990. How reliable are growth back-calculations based on otoliths? Can. J. Fish. Aquat. Sci. 47:2219-2227. Campana, S. E., and P. C. F. Hurley. 1989. An age- and temperature-mediated growth model for cod [Gadus morhua) and haddock {Melanogrammus aeglefmus) larvae in the Gulf of Maine. Can. J. Fish. Aquat. Sci. 46:603-613. Chambers, R. C. 1997. Environmental influences on egg and propagule sizes m marine fishes. Chapter 3 in R. C. Chambers and E. A. Trippel (eds.). Early life history and recruitment in fish populations, p. 63-102. Chapman and Hall, London. Chambers, R. C, and T. J. Miller. 1995. Evaluating fish growth by means of otolith increment analysis: special properties of individual-level longitudi- nal data. In D. H. Secor, J. M. Dean, and S. E. Campana (eds. I, Recent developments in otolith research, p 15.5- 175. Univ South Carolina Press, Columbia, SC. Crowder, L. B., J. A. Rice, T. J. Miller, and E. A. Marschall. 1992. Empirical and theoretical approaches to size-based interactions and recruitment variability in fishes. In D. L. DeAngelis and L. J. Gross (eds.), Individual-based ap- proaches in ecology, p 237-255. Chapman Hall, London. Francis, R. I. C. C. 1990. Back-calculation of fish length: a critical review. J. Fish Biol. 36:883-902. Fritz, E. S., L. B. Crowder, and R. C. Francis. 1990. The National Oceanic and Atmospheric Administra- tion plan for recruitment fisheries oceanography. Fish- eries 15:25-31. Geffen, A. J. 1995. Growth and otolith microstructure of cod (Gadus morhua L.) larvae. J. Plankton Res. 17:783-800. Hermann, A. J., S. Hinckley, B. A. Megrey, and P. J. Stabeno. 1996. Interannual variability of the early life history of walleye pollock near Shelikof Strait as inferred from a spa- tially explicit, individual-based model. Fish. Oceanogr. 5 (suppl. l):39-57. Knutsen, G. M., and S. Tilseth. 1985. Growth, development, and feeding success of Atlan- tic cod larvae Gadus morhua related to egg size. Trans. Am. Fish. Soc. 114:507-511. Lochmann, S. E., C. T. Taggart, D. A. Griffin, K. R. Thompson, and G. L. Maillet. 1997. Abundance and condition of larval cod (Gadus morhua) at a convergent front on Western Bank, Scotian Shelf Can. J. Fish. Aquat. Sci. 54:1461-1479. McLaren, I. A., and P. Avendaiio. 1995. Prey field and diet of larval cod on Western Bank, Scotian Shelf Can. J. Fish. Aquat. Sci. 52:448-463. Meekan, M. 1997. Relationship between otolith and somatic growth of cod larvae (Gadus mor/ii/a). J. Plankton Res. 19:167-169. Meekan, M. G., and L. Fortier. 1996. Selection for fast growth during the larval life of At- lantic cod Gadus morhua on the Scotian Shelf. Mar. Ecol. Prog. Ser. 137:25-37. Miller, T. J. 1997. The use of field studies to investigate selective pro- cesses in fish early life history. Chapter 7 in R. C. Cham- bers andE. A. Trippel (eds. I, Early life history and recruit- ment in fish populations, p. 197-223. Chapman and Hall, London. Miller, T. J., T. Herra, and W. C. Leggett. 1995. An individual-based analysis of the variability of eggs and their newly hatched larvae of Atlantic cod (Gadus morhua) on the Scotian Shelf Can. J. Fish. Aquat. Sci. 52:1083-1093. O'Boyle, R. N., M. Sinclair, R. J Conover, K. H. Mann, and A. C. Kohler. 1984. Temporal and spatial distribution of ichthyoplankton communities of the Scotian Shelf in relation to biological, hydrological, and physiographic features. Rapp. P-V. Reun. Cons. Int. Explor. Mer 183:27-40. Radtke, R. L. 1984. Cod fish otoliths: information storage structures. Flodevigen Rapportser 1:273-298. 1989. Larval fish age, growth, and body shrinkage: infor- mation available from otoliths. Can. J. Fish. Aquat. Sci. 46:1884-1894. Rice, J. A., L. B. Crowder, and E. A. Marschall. 1997. Predation on juvenile fishes: dynamics interactions between size-structured predators and prey. Chapter 12 in R. C. Chambers and E. A. Trippel (eds.). Early life his- tory and recruitment in fish populations, p. 333-356. Chapman and Hall, London. Ruzzante, D. E., C. T. Taggart, and D. Cook. 1996. Spatial and temporal variation in the genetic com- position of a larval cod (Gadus morhua) aggregation: co- hort contribution and genetic stability. Can. J. Fish. Aquat. Sci. 53:269.5-2705. Suthers, I. M., K. T. Frank, and S. E. Campana. 1989. Spatial comparison of recent growth in postlarval Atlantic cod(Gadus morhua ) off southwestern Nova Scotia: inferior growth in a presumed nursery area. Can. J. Fish. Aquat. Sci. 46:113-124. Taggart, C. T., K. R. Thompson, G. L. Maillet, S. E. Lochmann, and D. A. Griffin. 1996. Abundance distribution of larval cod (Gadus morhua) and zooplankton in a gyre-like water mass on the Scotian Shelf In Y. Watanabe, Y. Yamashita, and Y. Oozeki (eds.). Survival strategies in early life stages of marine resources, p. 15.5-173. Balkema, Rotterdam. Thompson, B. M., and J. D. Riley. 1981. Egg and larval development studies in the North Sea cod (Gadus morhua L.). Rapp. P-V. Reun. Cons. Int. Explor Mer 178:553-559. 306 Abstract.— Daily growth increments on otoliths were used to estimate the age of larval and juvenile haddock, Melanogrammus aeglefinus. and pol- lock, Pollachius virens, collected on Emerald and Sable Island Banks, east- ern Canada, between March 1991 and May 1993. The daily periodicity of the increments was validated from obser- vations of reared larvae. For both spe- cies, the first increment was deposited the day after hatching and thereafter one increment was added daily. A Laird- Gompertz growth curve was fitted to length-age data for each species. Growth rates in haddock and pollock larvae varied significantly in different years. For haddock, the lowest growth rate was for the 1993 cohort, and growth rates in 1991 and 1992 cohorts were similar. For pollock, the 1993 co- hort had the highest growth rate. The average growth rate was 0.21 mm/d for the first month and 0.42 mm/d for the second month for larval haddock and 0.18 mm/d for the first month and 0.23 mm/d for the second month for larval pollock. Growth continued exponen- tially after the transition from a pri- marily pelagic life to a predominantly demersal one, which occurred at an age of about 40-.50 d. No indication of a ces- sation in growth was observed. Analy- sis of length-age data indicated that the accelerated growth of juveniles af- ter 50 d in age could have reflected the exploitation of a more abundant food resource after settlement. Thus, pelagic and early demersal growth appear to represent distinct stanzas in the growth history of these gadoids. Age validation and growth of larval and juvenile haddock, Melanogrammus aeglefmus, and pollock, Pollachius virens, on the Scotian Shelf Casimiro Quiiionez-Velazquez Centre Interdisciplinano de Ciencias Marinas Playa El Conchalito s/n PO Box 592 La Paz, BCS, Mexico E-mail address cquinone a vmredipn ipn mx Manuscript accepted 24 June 1998. Fish. Bull. 97:306-319 ( 1999). A central problem in fisheries re- search is understanding mechanisms determining year-class strength. Growth, which is considered critical to the survival and subsequent re- cruitment of larval marine fishes (Houde, 1987), is likely strongly af- fected by temperature and food availability (Ricker, 1979). Food limitation can affect survival di- rectly by causing starvation (Lasker, 1975) or indirectly by retarding growth rate which in turn increases mortality from predators (Roths- child and Rooth, 1982). In addition, feeding conditions may change markedly owing to density indepen- dent factors, such as temperature, which can directly affect growth rates (Laurence, 1978). Ware ( 1975 ) suggested that growth and mortality rates interact to deter- mine survival offish populations, A prediction of this hypothesis is that predation is a major factor affect- ing year-class strength and that mortality due to predation is in- versely related to growth rate (Ware, 1975; Shepherd and Gushing, 1980). This hypothesis assumes that average growth is below the maxi- mum and that feeding conditions that maximize growth are associ- ated with minimal mortality. Galculation of reliable rates of growth and mortality of larval fish, and determination of when loss due to recruitment is greatest, requires accurate determinations of age. Ac- curacy and precision of growth es- timates for larval fishes have been greatly enhanced by the discovery of daily growth increments on otoliths (Pannella, 1971; reviewed by Gampana and Neilson, 1985). Ageing by counting otolith growth increments, along with measure- ments of the width of the incre- ments, provides a means of estimat- ing the relation between length at age. In this way, growth curves, and even individual growth rates, have been calculated for a variety of spe- cies (Brothers et al,, 1976; Struh- saker and Uchiyama, 1976;Taubert and Goble, 1977; Radtke and Wai- wood, 1980). Daily increments have been shown to occur in most species, nev- ertheless validation of the fre- quency of increment deposition is required before larval age can be estimated from otolith measure- ments. To validate increment depo- sition, Geffen (1987) recommended sampling reared larvae throughout the larval stage. This approach had previously been used for a number of species, for example, to determine when the first increment is depos- ited in Atlantic cod (Gadus morhua) larvae ( Radtke and Waiwood, 1980) and to validate the daily increment formation in Pacific herring (CY (/pea pallasi) (Moksness and Wespestad, 1989). In the present study, I vali- Quinonez-Velazquez: Age validation and growth of Melanogrammus aeglefinus and Pollachius virens 307 dated whether increments are deposited daily in haddock otoUths (Melanogrammus aeglefinus ) using the same approach. PreHminary results on larval pol- lock (Pollack ius virens ) are also presented. Finally, daily increment analyses are used to estimate age and growth rates of larval and juvenile haddock (M. aeglefinus) and pollock (P. virens) from the Emerald and Sable Island Banks on the Scotian Shelf off eastern Canada. Materials and methods Study area and collection of larvae The Scotian Shelf is a 62,000 km^ area with an aver- age depth of 90 m. It is bounded on the southwest by Northeast Channel entering the Gulf of Maine and on the northeast by the Laurentian Channel and Gulf of St. Lawrence. Several shallow offshore banks along the outer edge of the shelf are separated from one another by deeper channels that open onto interior basins having depths of >100 m (O'Boyle et al., 1984). The general water circulation on the Scotian Shelf is dominated by the Scotian current which flows south- westerly and parallel to the Nova Scotia coast and which carries a mixture of slope water and diluted waters from the Gulf of St. Lawrence (e.g. Sutcliffe et al.. 1976; Drinkwater et al., 1979). Eggs, larvae, and juveniles of M. aeglefinus and P. virens were collected during 20 surveys made on Emerald and Sable Island Banks (Fig. 1) from March 1991 to May 1993 (Table 1), as part of the program of Ocean Production Enhancement Network (OPEN) (Griffin and Anderson, 1993). Sampling was con- ducted using a grid of 45 stations (Fig. 1) with an interstation spacing of approximately 30 km. A 50- cm bongo frame with 150- and 250-|.im mesh nets (March-May 1991 ) as well as a rectangular midwater trawl (RMT, 2-8 m^) (Baker et al., 1973) with nets of 2 m^ (333-|.im mesh) and 8 m- (1600-|.tm mesh) were used. At each station, duplicate oblique hauls were made from near the bottom to the surface at 2.0-2.5 knots, with a retrieval time of 30 min. After the stan- dard grid stations were sampled, the area of maxi- mum abundance of larval fish was revisited so that 48 h of vertical sampling could be conducted with a BIGNESS sampler (1 m-) (Eastern Marine Services E-Z-Net, Dartmouth, NS) fitted with ten 333-nm mesh nets (June 1991 to May 1993). Ten discrete depths were sampled every 4 h. Deployment sam- pling depths were spaced at 5-m intervals, from the bottom to the surface, and fished for 5 min at 2.0- 2.5 knots. Most larval haddock and pollock were - sorted at sea and preserved in 95% ethanol. Tempera- ture and salinity profiles were recorded at each station Figure 1 Study area showing stations where lar\'al and juvenile had- dock and pollock were collected on the Scotian Shelf from March 1991 to May 1993. Isobaths are in meters. with a Sea Bird CTD probe (Sea-Bird Electronics, Inc. Bellevue, WA ) connected to the electronics unit of the RMT or BIGNESS sampler, or both. When the RMT or BIGNESS, or both, were not used, independent CTD casts were made at each station. A subset of pollock and haddock larvae were vid- eotaped at sea by using a stereo dissecting micro- scope connected to a video camera and a magneto- scope and were then individually preserved in 95% ethanol. One to two months later, each fish was mea- sured in the laboratory to the nearest 0.1 mm with a stereo dissecting microscope with an ocular microme- ter. The unpreserved standard lengths were mea- sured from video images recorded at sea and pro- jected on a television monitor linked to an image analysis system. Linear regression of unpreserved standard length (SL, obtained from the videotape) to preserved standard length (PSL. after preserva- tion in ethanol) for haddock (SL=1.531 -i- 1.005PSL, /! = 129, r-=0.99) and pollock (SL=0.710 -i- 0.969PSL, 308 Fishery Bulletin 97(2), 1999 Table 1 Station information for larval and juveni e of haddock and pollock cc Uected for otolith analysis from March 1991 to May 1993 on the Scotian Shelf. Values in parentheses are total larvae sampled. No. of larvae Mean SL of Mean no. of examined larvae (mm) increments T (°C) Cruise Vessel Date at 30 m Haddock Pollock Haddock Pollock Haddock Pollock 91-01 Petrel V 5-11 Mar 4.0 0(3) 3.9 91-02 Petrel V 16-24 Apr 5.0 2(97) 4.6 14 91-04 Petrel V 17-26 May 7.0 1 (221 7.6 23 91-99 Cape Keltic 17-22 Jun 8.3 93 1164) 20.2 60 91-08 Petrel V 18-27 Jul 13.5 18(181 60.8 105 91-16 Petrel V 12-21 Nov 8.8 4(16) 3.9 4 91-17 Petrel V 7-16 Dec 7.3 12(23) 5.1 9 92-18 Petrel V 7-18 Jan 5.2 88(98) 5.7 11 92-19 Petrel V 16-26 Feb 2.3 0(9) 80(218) 5.1 10.7 29 92-20 Petrel V 10-20 Mar 1.6 12 (66) 17(61) 4.9 12 2 41 92-21 Petrel V 5-14 Apr 2.0 9(40) 7 (33) 4.7 15 3 44 92-23 Petrel V 17-26 May 3.8 65 (328) 2(2) 8.1 22 21 70 92-98 Cape Keltic 17-22 Jun 7.3 16(47) 22 60 92-26 Petrel V 21-25 Jul 13.7 5(5) 31.1 66 93-32 Petrel V 11-22 Jan 4.0 246(1005) 5.4 5 93-33 Petrel V 15-25 Feb 2.1 54(57) 9.9 27 93-34 Petrel V 19-26 Mar 1.7 138(266) 5)5) 4.4 15.1 4 54 93-35 Petrel V 16-24 Apr 3.7 116(239) 4)4) 7.2 13.5 16 48 93-37 Petrel V 16-24 May 5.6 127(290) 0 (10) 12.4 15.5 33 n=8, r^=0.99) were used to calculate the unpresei-ved standard length (SL) offish preserved in ethanol. Validation of ageing technique To determine if the increments observed in otoliths of haddock and pollock were deposited daily, fertil- ized eggs collected in a 333 yim mesh net from the RMT were reared on the ship. Haddock eggs were collected at the Western and Emerald Banks area, February-April 1992 and 1993. Spawning begins in February around the Western Bank. Eggs were taken in the same general area on any given sampling date from February to April in both 1992 and 1993. Some pollock eggs were collected in January-Feb- ruary and April 1992, as well as in October-Novem- ber 1992 and January and March 1993. At the be- ginning of the spawning season, eggs were detected at the southeast of Sable Island Bank. In January- February the eggs were collected on the Western Bank, and in March-April the distribution of eggs included Emerald Bank. The spawning of pollock and haddock overlapped during February-April because the eggs of the two species were found together. Immediately after the nets were brought aboard the vessel, the most advanced stages of eggs were sorted and individually placed in 20-mm vials filled with seawater which had been passed through l-|im Hytrex filters. At this time it was impossible to dis- tinguish between the eggs of the two species (Brander and Hurley, 1992). They were reared at 4°C in a con- ventional refrigerator on the ship. The photoperiod was kept at 12:12 h with dim blue light. As each egg was monitored twice daily to establish when hatch- ing occurred, the maximum error in hatching time determinations was 12 h. After hatching, 136 zero to eleven-day haddock larvae, and nine zero to four-day pollock larvae were killed and preserved in 959i etha- nol. Twelve additional 0-2 d old haddock larvae (col- lected on April 1993) were obtained from Flodevigen Marine Research Station, in Arendal, Norway. A to- tal of 148 haddock larvae and 9 pollock larvae were used to examine increment formation. Because had- dock and pollock eggs were indistinguishable; individu- als were later identified by their larval characteristics. Otolith preparation A dissecting microscope illuminated with polarized light was used to dissect the sagittae and lapilli from larval skulls with the aid of fine needles. One thou- sand two hundred and sixty pairs of haddock and Quihonez-Velazquez: Age validation and growth of Melanogrammus aeglefinus and Po/lachius virens 309 pollock larval otoliths from the validation experiment and from wild fish <36 mm SL were mounted on glass over slides with Permount (Fisher Scientific Labo- ratory, Fair Lawn, NJ). Eighteen pairs of haddock otoliths from larvae measuring 50-83 mm SL were mounted with synthetic resin on metal stubs for SEM observation. The mounted otoliths (sagittae) used in the vali- dation experiment were read with an image analyzer system by using Optimas software (Subtechnique, Inc., Alexandria, VA). Otoliths from field-collected larvae and juveniles were read at the Southwest Fish- eries Center (La Jolla, CA) by using the OTO pro- gram (Andersen and Mokness, 1988). The image analyzer system consisted of a video camera attached to a compound microscope, monitor, digitizer, and microcomputer. A detailed description of the OTO program, otolith analyzing system, and methods used are given by Andersen and Moksness ( 1988 ). Otoliths examined with SEM were ground on the sagittal plane to the nucleus with fine grit paper (between 30 |am and 0.3 |am), and then the polished surface was etched for 6 min in 0.2M EDTA (pH 7.6). In addition to the number of increments, the fol- lowing otolith measurements were taken and used in comparisons of each species: maximum otolith di- ameter (OD); maximum otolith radius from the cen- ter of the nucleus to the outer edge of the otolith (OR); diameter of the otolith nucleus (ON); and diameter of the yolksac resorption check (YSC) as defined by Radtke and Waiwood (1980) and Bolz and Lough (1983). Growth increments were counted twice by the same reader at an interval of 3 mo between read- ings. The two counts usually differed <69'f, and the average was used as an estimate of age. Statistical procedures For both species, of all otolith measurements, the maximum otolith diameter was most strongly corre- lated with length. I compared the relation of otolith growth to larval growth between month and year for both species to determine whether otolith growth could be used to predict lai-val growth. Thus, linear regres- sions of LnSL on LnOD were compared by using analy- sis of covariance ( ANCOVA) with OD as the covariate. A Laird-Gompertz growth model curve was fitted to the length-at-age data for each species in each year. This model has been shown to provide an adequate fit for length-at-age data on age 0+ fish of many dif- ferent species (Bolz and Lough, 1988; Lough et al., 1982; Watanabe et al., 1988; Simard et al., 1992). Zweifel and Lasker ( 1976) presented a detailed dis- ' cussion of the Laird-Gompertz function. The equa- tion for the model is Lt L^e kd-e"") where L„ L. - "0 = length at ^=0; a dimensionless parameter, such that ka=AQ is the specific growth rate at /=0 (A,=Aoe^"; length at any age t; the specific rate of growth when t-t^^; and the time when the growth rate starts to decrease, that is, the inflection point of the curve (Ricker, 1979). The parameters were derived by nonlinear least squares tests by using the SYSTAT nonlinear pro- gram (Wilkinson, 1990), and ANCOVA was used to test whether differences in growth rates in different years of each species were significant. Results Validation of daily increment deposition In both species, otoliths (sagittae and lapilli) were present at hatching. Some haddock otoliths ( Fig. 2A) showed one to three irregular increments between the nucleus and the check at hatching, as observed by Bolz and Lough (1983). The mean (±SD) dia- meters of the nucleus and of the check at hatching of reared larvae were 10.8 ±2.5 and 21.4 ±3.3 |am, re- spectively, in haddock and 12.0 ±1.9 and 20.7 ±2.2 Hm, respectively, in pollock. The values for haddock were consistent with earlier reports by Bolz and Lough (1983) and Campana (1989). In both species, increments appeared as alternate light and dark zones (Fig. 2), and the first regular increment was formed the day after hatching (Table 2). The num- ber of increments (NI) corresponded to the chrono- logical age in d ( AGE ) of the larvae ( Fig. 3 ) . The slopes of the regressions for haddock {NI-0.99AGE, /? = 148, r~=0.96, P>0.0001) and pollock {NI=l.09AGE, n=9, r''^0.82, P=0.0005 ) did not differ significantly from 1 (^test, P=0.637 for haddock and P=0.890 for pollock). Growth of haddock and pollock otoliths The check at hatching was clearly visible and incre- ments were easily distinguished in the otoliths of larvae sampled at sea (Fig. 4). The nuclear and check diameters at hatching for haddock and pollock dif- fered in different years (AN OVA, P<0.01) (Table 3). For haddock, differences in nuclear check diameters between 1991 and 1993 were not significant (AN OVA, P>0.05). There were no significant differences among 310 Fishery Bulletin 97(2), 1999 years for yolksac check diameters for either haddock or pollock larvae (ANOVA, P>0.0.5). Within some haddock otolith nuclei, 1-3 increments were ob- served. Outside the nucleus, 6-8 growth increments of irregular width were observed between the nucleus check and a second well-marked check. Bolz and Lough ( 1983 ) observed 2-8 faint increments bounded by a discontinuous zone that appeared to be analo- gous to the yolksac check found in larval cod otoliths by Radtke and Waiwood ( 1980). Accordingly, I inter- preted the second well-defined mark as a check resulting from yolksac resorption. The mean dia- meter of the yolksac check was 34.9 ±2.8 |im in had- dock and 35.3 ±2.6 ^m in pollock. Several (mean 7 ± 1.8) narrow (<1 |im) increments were found immedi- ately after the yolksac resorption check. Increment width increased rapidly in subsequent increments. The relation between otolith and lar- val fish sizes was best described by a In- In regression (Fig. 5). For haddock and pollock, significant differences were found between cruises ( ANCOVA, P<0.05 ); how- ever, no significant differences were found among years (ANCOVA, P>0.05) for either species (Table 4). Among had- dock samples, a multiple range test re- vealed four homogeneous groupings in the samples from the ten cruises analyzed (P<0.01). The groupings were interpret- able in terms of monthly sagittal growth. The first group integrated larvae from March to April; the second, larvae from May; the third, larvae from June; and the fourth, larvae from July. With pollock, the multiple range test showed three homo- geneous groupings among the pollock sampled during the ten cruises (P<0.01 ). The first group integrated larvae from November to January; the second, larvae from February; and the third, larvae from March to April. It was possible to distin- guish different stages in sagittal growth from otolith structure. Sagittae from the early developmental stages of haddock and pollock larvae were almost circular in shape, and there was one fiat and one convex-side, accessory nucleus developed in the sagittae of postlarvae. As the lar- vae developed, the otoliths grew faster along the anteroposterior axis and be- came oval shaped. Haddock and pollock larval and juvenile growth Figure 2 Light micrographs of sagittae. (A) Sagitta from a ,5.93-mm, .5-day-old, haddock lar\'a. (B) Saggita from a'4.98-mm, 3-day-old, pollock lari'a. H = hatch check. Growth increments appear as light and dark zones. The lan'ae were hatched from eggs collected from the Western and Emerald Banks area. Bar represents .5 [jm. For both haddock and pollock, the first otolith increment was laid down the day after hatching. Thus, the age of larvae and juveniles was indicated by the num- ber of increments in the sagittae. Daily increments were counted in otoliths from 1121 individuals of a total of 3126 cap- Quinonez-Velazquez: Age validation and growth of Melanogrammus aeglefinus and Pollachius virens 311 Table 2 Chronological age (from hatching) otolith diameter, and incremer t formation in reared 1 irval haddock and pollock. Chronological age (days) Standard length (mm) Otolith diameter (|im) Increment ±SD Number of larvae Haddock 0 4.69 26.70 0.5 0.5 18 1 4.75 27.87 1.1 0.2 15 2 5.12 30.04 2.1 0.3 21 3 5.16 32.11 3 0.4 15 4 5.68 33.03 4.1 0.4 26 5 5.74 34.03 5.1 0.3 13 6 5.86 34.04 5.8 0.6 23 7 5.90 34.41 6.6 0.8 9 8 5.86 36.45 7.3 0.4 4 9 5.77 36.92 8.5 0.5 2 10 5.84 37.58 10 — 1 11 6.19 37.67 11 — 1 Pollock 0 4.10 24.72 1 1 1 3.81 26.10 1 — 1 2 4.34 29.38 2 0 3 3 5.00 28.78 3.3 0.5 3 4 4.49 32.49 4 — 1 tured. Haddock larvae and juveniles («=602) ranged from 2.7 to 83.0 mm in SL and from 0 to 128 d in age, and pollock larvae and juveniles (n=519) from 2.8 to 23.8 mm in SL and from 0 to 84 d in age. For each year, Gompertz growth curves were fitted to describe the mean growth of larval and juvenile haddock (1991, 1992, and 1993) and pollock (1992 and 1993) (Fig. 6; Table 5). Length at hatching predicted from the curves was 4.1 mm for haddock and 4.5 mm for pollock, both within the range reported by previous studies (Fridgeirsson, 1978; Bolz and Lough, 1983; Fahay, 1983). Because the predicted inflection point for both haddock and pollock curves fell beyond the length range analyzed, the end of the first growing season had not been reached. Although haddock larvae older than 80 d were ob- tained in 1991 and 1992, they were not used in the comparison of growth. Haddock growth rates (mm/d) during the first 80-d period in 1993 were different from those in 1991, and growth rates in 1993 were different from those in 1992 (ANCOVA, P<0.01) (Table 6 ). The 1993 cohort had the lowest growth rate. The 1991 and 1992 cohorts had similar growth rates (ANCOVA, P=:0.84). The average growth rate of lar- vae was 0.21 mm/d during the first month and 0.42 ~mm/d during the second month. Growth continued exponentially from the predicted length at hatching Table 3 ANOVAo f nucleus, hatching. and yolksac checks on year for haddock and pollock larvae collected on the Scotian Shelf. 1991 1992 1993 F P>F Haddock Nucleus' 15.7 16.9 15.5 40.8 0.00 SD 1.4 1.3 1.4 Hatching 22.4 25.1 23.5 34.9 0.00 SD 2.6 3 2.1 Yolksac 36.6 38.7 35 0.28 0.76 SD 3.9 4.1 2.7 Pollock Nucleus SD 14.8 1.6 15.9 1.8 38.7 0.00 Hatching 23.9 23.1 38.1 0.00 SD 1.4 1.9 Yolksac 35.6 35.1 5.2 0.02 SD 2.6 2.7 ' Nucleus check 1991 versus 1993 not significant (/'>0.42) throughout the size range of larvae and juvenile col- lected, there being no indication of a cessation in growth. Pollock growth rates (mm/d) varied significantly among years (ANCOVA, P<0.01 ) (Table 6). However, 312 Fishery Bulletin 97(2), 1999 12 o O 12 10 ■ 8 ■ 6 4 2 ■ 0 - PoUachius virens 2 4 6 8 10 Age (days after hatching) 12 Figure 3 Number of increments enumerated on the sagittae of haddock and pollock larvae as a function of known age (days after hatching). Numbers represent the number of larvae. The number of increments (NI) cor- responds to the chronological age in d (AGE) of the lar%'ae. Haddock NI = 0.99 AGE. n = 148. r- = 0.96. and pollock NI = 1.09 AG£, n = 9, /-' = 0.82. the Aq |Specific growth rate at ^=0) did not vary sig- nificantly (ANCOVA, P>0.05). The 1992 cohort had the higher growth rates during the first 50 d follow- ing hatching. Growth was linear at 0.18 mm/d fol- lowing the predicted hatching size (1992 cohort), 0.13 mm/d (1993 cohort) through the first month, and 0.23 (1992 cohort) and 0.24 mm/d (1993 cohort) through the second month. Discussion and conclusions Validation of increment deposition The present study shows that growth increments in haddock otoliths are formed daily and suggests that this is also the case in pollock. In both species, the first of the regular increments is formed on the day after hatching, and thereafter an additional incre- ment is added each day up to 1 1 d for haddock and 4 d for pollock. Otoliths develop before hatching, and in haddock, 1 to 3 irregular increments are depos- ited prior to hatching. These types of irregular in- crements were not observed on larval otoliths of pol- lock. A few embryonic growth increments of a lamel- lar structure (such as I observed in haddock), have also been reported in other species (Brothers et al. , 1976; Radtke and Dean, 1982; Watanabe et al. .1982; Nishimura andYamada, 1984;Palomeraet al., 1988). The spacing of embryonic increments is irregular, in contrast to regularly spaced posthatching incre- ments, which indicates that they might not be de- posited at daily intervals. Whereas Radtke and Waiwood (1980) did not observe increments formed prior to hatching in cod (G. morhua), they did ob- serve regular increments that were formed daily starting one day after hatching. Brothers and McFarland ( 1981 ) noted three diffuse increments in the otolith core of French grunt (Haemulon flavo- lineatiim ) and speculated that they were deposited during the embryonic stage. Bolz and Lough (1983) found 1-2 poorly defined increments enclosed by the nuclear check in haddock larvae. According to the above observations, it seems reasonable to infer that the 1-3 irregular increments inside the hatch check of haddock otoliths are formed sometime during the 2-3 week egg stage (Laurence and Rogers, 1976; Fridgeirsson, 1978). Growth of haddock and pollock otoliths The sagittal otoliths of haddock and pollock larvae are circular in shape at the age of 35-40 d. Subse- quent sagittal growth is greater along the antero- posterior plane, so that the otoliths become oval shaped in older larvae. Radtke ( 1989) reported that cod (G. morhua ) sagittae change as they grow, from spherical to elongated to crenulated over a 65-d pe- riod. In haddock and pollock larvae, circular-shaped sagittae were observed until 8 mm SL, correspond- ing to an age of 35-40 d. Then, the sagittal shape begins to elongate as the result of growth of periph- eral primordia, which corresponds to 30 mm SL (75- 80 d) in haddock and 23 mm SL (80 d) in pollock. Thereafter the sagittae are not easily broken, and their shape is distinguishable from that of other ga- doid species. The last changes in otolith shape corre- spond to the age when juveniles undergo metamor- phosis to begin demersal life. Koeller et al. (1986) inferred, from a series of midwater and botton trawl surveys conducted in 1983 off southwestern Nova Scotia, that the transition to the demersal habitat by haddock occurs from June to August. In most otoliths examined, I observed a well-de- fined, dark and discontinuous zone or ring laid down Quinonez-Velazquez: Age validation and growth of Melanogrammus aeglefinus and Pollachius virens 313 A C 4» « 4 .J' YS V I " Ik B ^Pp i^ YS X Figure 4 Light micrographs of sagittae showing different landmark checks. (A) Sagitta from a 13.7-mm-SL, 37-day-old haddock. (B) Sagitta from a 12.4-mm-SL, 50-day-old pollock N = nuclear check, H = check at hatching, and YS = yolksac check. Bar of photograph represents 10 |am. around the nucleus. It was the first mark presented in larval otoliths (Bolz and Lough, 1983; Radtke and Waiwood, 1980; Campana. 1989). Outside the nucleus, four to eight irregularly spaced increments were bounded by a discontinuous zone which ap- peared to be the yolksac check (Bolz and Lough, 1983). Laurence ( 1974) reported that at 7^C yolksac exhaustion occurs 6-7 d after hatching for haddock, and Fridgeirsson (1978) observed that at 7.2°C yolksac exhaustion occurs within 6 d for haddock larvae and 8 d for pollock larvae. From the regres- sion of larval length to sagittal diameter, I estimated that yolksac resorption occurs at a standard length of 5.0 mm in haddock (at 3.5°C) and 5.1 mm in pol- lock larvae (at 5.0°C), corresponding to a mean age of 9 and 5 d, respectively. Fridgeirsson ( 1978) reported that haddock start independent feeding at 6-10 d after hatching and pollock at 8-10 d after hatching. 314 Fishery Bulletin 97(2), 1999 This suggests that the observed check is deposited at or immediately before yolksac absorption and that the number of increments between the nucleus and the absorption check reflects the duration of the yolksac phase (Fridgeirsson, 1978; Laurence, 1978; Fahay, 1983). Haddock and pollock larval and juvenile growth In recent years, the examination of otolith micro- structure has resulted in many applications (Jones, 1992). The underlying assumption is that increments form daily. In otoliths of haddock (A/, aeglefinus ), and pollock {P. virens). increments are deposited daily and thus can provide accurate estimates of age (Bolz and Lough, 1983; Campana, 1989; Campana and Hurley, 1989). The estimates of haddock growth in this study are comparable with those reported earlier. Thus, the predicted length of 4.1 mm at hatching and the av- erage growth rate of 0.65 mm/d for 0-4 mo old lar- vae in the present study are similar to the values reported by Bolz and Lough (1983, 1988). The ad- justment of a Laird-Gompertz model (/■"=0.99) for the InSI =0 571nOD-0 48 F= 14689. PO «=471. R=0 98 1,000 2,000 3,000 4,000 5,000 30 25 20 15 10 5 0 100 200 300 400 500 600 Sagittal maximum diameter (^m) Figure 5 Standard length as a function of ma.ximum .sa^jiltal diameter in larval pollock and haddock. PoUachius virens t^ oo r^ fo^ lnSL=0 50lnOD-0 15 F=336l. POOOOI n=532, R=0 93 entire data set correctly predicted length (Table 5). Different patterns in growth occur during ontogeny (Table 6). The average increase in length is moder- ate between hatching and 50 d (0.27 mm/d), high for early juveniles of 2.5-3 mo (0.68 mm/d), and moder- ate again (1.2 mm/d) for older juvenile. Few studies provide estimates of the age and growth of larval and juvenile pollock (Campana, 1989). The predicted length at hatching was 4.5 mm and the average growth rate of larval was 0.22 mm/d during the first 3 mo. The Laird-Gompertz model (r-=0.98) applied to the entire set data adequately described growth. A slow increase in length occurs during the first 30 d after hatching (0.18 mm/d ), and then the rate is mod- erate during the following 1 to 2 mo (0.23 mm/d). Finally, a stable period (0.24 mm/d) begins (Table 6). In the present study only the pelagic phase prior to metamorphosis and beginning of demersal life was examined. Nevertheless, sampling continued throughout the year and older pollock juveniles up to 80 d old (23 mm SL) were captured. The last pol- lock samples were collected in 1993. Juvenile pol- lock probably move from the oceanic zone to the shal- low coastal zone after metamorphosis (Scott and Scott, 1988). Campana (1989) collected 53 pol- lock juveniles from the north shore of Gran Manan Island, New Brunswick, on August 1984, and the mean length of juveniles was 97.3 ± 1.9 mm. The date of capture of Campana's samples suggest that pollock juvenile move to the Scotian shore, possibly toward the Bay of Fundy, to fin- ish their development. If pollock sharply increase their growth rate in response to improved feed- ing conditions during demersal lifestyle, as we observed for haddock, they could attain an aver- age length of 106 mm by August, a size similar to that of juveniles collected by Campana. Growth rates (mm/d) varied significantly in different years, and the rates of the two cohorts showed an alternate pattern: growth was higher for the 1992 cohort during the first 45 d and higher for 1993 cohort following the first 45 d. The 1993 cohort was first detected two months later than the 1992 cohort. The adjustment of a Laird- Gompertz model (r^=0.98) to each annual data set closely predicted length (Table 5). Growth during the juvenile period of haddock and pollock can be divided into a series of "stan- zas." The change from one stanza to the next is characterized by a discontinuity in development, such as during hatching or after a change of habi- tat (Ricker, 1979). Growth accelerates abruptly in fish older than 50 d in response to a shift from a pelagic to a demersal diet. Because we captured juvenile pollock immediately after the transition Quinonez-Velazquez: Age validation and growth of Melanogrammus aeglefinus and Pollachius virens 315 Table 4 Parameter est mates of the relationship between standard length and sagittal diameter for larval haddock and pollock collected on the Scotian Shelf. LnSL = a + 6LnSD. Note: The reduced correlation during March 1992 and 1993 for larval haddock is the | result of differences between size of larvae and sagittal diameter at hatching. Species Year Month Intercept a) Slope (6) n r2 Estimate SE Estimate SE Haddock 1991 June 0.006 0.090 0.477 0.014 37 0.97 July -3.140 0.442 0.898 0.055 18 0.94 1991 -0.524 0.077 0.567 0.011 55 0.98 1992 March 0.840 0.781 0.157 0.216 12 0.05 April -0.998 0.422 0.705 0.120 12 0.77 May 0.068 0.101 0.466 0.022 65 0.87 June -0.002 0.167 0.500 0.027 16 0.96 July -0.722 0.530 0.594 0.078 5 0.95 1992 -0.466 0.059 0.572 0.012 110 0.95 1993 March -4.574 0.467 1.733 0.134 134 0.56 April -0.618 0.162 0.612 0.038 45 0.86 May 0.047 0.056 0.466 0.011 127 0.94 1993 -0.469 0.032 0.564 0.007 306 0.95 All -0.445 0.021 0.560 0.004 471 0.97 Pollock 1991 November 2.162 0.239 -0.224 0.069 4 0.84 December -0.646 0.409 0.613 0.113 12 0.75 1992 January -0.943 0.154 0.717 0.041 91 0.77 February -0.095 0.093 0.482 0.019 80 0.89 March -1.011 0.325 0.644 0.060 17 0.88 April -0.463 0.469 0.568 0.087 7 0.90 1991- 1992 -0.143 0.045 0.494 0.010 211 0.92 1993 January -0.932 0.132 0.720 0.036 258 0.60 February 0.694 0.119 0.326 0.024 54 0.78 March -0.856 0.334 0.632 0.060 5 0.97 April -0.392 0.505 0.538 0.091 4 0.95 1993 -0.204 0.057 0.517 0.015 321 0.80 All -0.143 0.036 0.498 0.009 532 0.86 to a demersal life stage, it was possible that indi- viduals were present in both pelagic and demersal zones at this time. Perhaps pollock juveniles were moving from the oceanic to the coastal zone; if so, this change in habitat could explain the reduction in numbers of juvenile older than 60 d in our sampling. I assume that conclusions drawn for juvenile had- dock are valid only during the first eighty days for pollock. This period corresponds to the transition from a primarily pelagic to predominantly demersal life stage, which occurs when haddock reach about 20-25 mm in length (Koeller et al., 1986) and when pollock reach 10-15 mm. Although >25-mm haddock were captured in the water column, they probably had adopted a predominantly demersal life style. Thus, the acceleration in growth of >50-d juveniles could have reflected exploitation of abundant food resource after settling on the bottom (Mahon and Neilson, 1987). In the present study, average growth rates were 0.21 mm/d in the first month and 0.42 mm/d in the Table 5 Estimates of Gompertz model parameters for larval and juvenile haddock and pollock collected on the Scotian Shelf Lp= lenght at age t = 0,a = specific rate of growth aXt = ?(,, and A=dimensionless parameter Species Year ^0 a k n r2 Haddock 1991 2.478 0.010 4.954 114 0.99 1992 3.609 0.006 6.172 107 0.99 1993 4.317 0.005 5.809 381 0.99 All 4.198 0.005 6.581 602 0.99 Pollock 1991-92 4.213 0.013 2.483 210 0.99 1993 4.617 0.017 1.970 309 0.98 All 4.569 0.013 2.429 519 0.98 second month for larval haddock and 0.18 mm/d in the first month and 0.23 mm/d in the second month for larval pollock. Growth continued exponentially 316 Fishery Bulletin 97(2), 1999 90 Melanogrammus aeglefimis 60 ■ 30 90 1 90 1 60 30 1991 1 1 1 1 1 50 100 150 ;^ 1993 Pollachius virens Age (days) 0 50 100 150 Age (days) Figure 6 Estimated age versus length for larval and juvenile pollock and had- dock. Lines were fitted by using the Laird-Gompertz growth model. Parameters of the model are shown in Table 5. from the predicted length at hatching throughout the size range of larvae and juvenile collected. Otolith growth increments are suitable for estimating age, and the analysis of length-age data for young had- dock and pollock on Emerald and Sable Island Banks suggests that the growth rate increases sharply at the transition from a pelagic to demersal life stage. Thus, pelagic and early demersal growth likely rep- resents distinct "stanzas" in the growth history of these gadoids, and these "stanzas" may have differ- ent effects on recruitment variability. Rapid juvenile growth may increase juvenile survivorship by de- creasing the time during which juvenile are exposed to predators (Houde, 1987). Interannual variation in juvenile growth rates may influence interannual variation in juvenile survivorship. Additional com- plicating factors during the pelagic phase of juvenile fish, such as different residence times and diel mi- grations in the water column (Koeller et al., 1986), reduce the possibility of identifmg correlations be- tween interannual variation in growth rates and environmental factors, and thereby allow predictions of recruitment success. Acknowledgments This study was supported by funding from the Natu- ral Sciences and Engineering Research Council of Canada (NSERC) to the Ocean Production Enhance- ment Network (OPEN) program. Support from Instituto Politecnico Nacional, COFAA, and CoNaCyT- Mexico was also provided to the author. I am most grateful to numerous colleagues from Dalhousie, Quiiionez-Velazquez: Age validation and growth of Melanogrammus aeglefinus and Pollachius virens 317 Table 6 Growth rate (mm/d and %/d) of larval and juvenile haddock and pollock estimated from the Laird-Gompertz growth model. Haddock Pollock Growth rates (mm/d) Growth rates (%/d) Growth rates (mm/d) Growth rates {%/d) Age (days) 1991 1992 1993 All 1991 1992 1993 All 1992 1993 All 1992 1993 All 0 0.13 0.13 0.11 0.14 4.00 3.70 2.56 3.85 0.14 0.10 0.14 3.23 1.85 3.16 5 0.16 0.16 0.12 0.16 3.86 3.59 2.56 3.72 0.15 0.10 0.16 3.02 1.85 2.96 10 0.18 0.18 0.14 0.18 3.73 3.49 2.56 3.59 0.16 0.11 0.17 2.83 1.85 2.77 15 0.21 0.21 0.16 0.21 3.60 3.38 2.56 3.47 0.17 0.13 0.18 2.66 1.84 2.60 20 0.24 0.24 0.18 0.24 3.48 3.28 2.55 3.35 0.19 0.14 0.19 2.49 1.84 2.43 25 0.28 0.27 0.21 0.27 3.36 3.19 2.55 3.23 0.20 0.15 0.20 2.33 1.84 2.28 30 0.32 0.31 0.24 0.31 3.24 3.09 2.55 3.12 0.21 0.16 0.21 2.19 1.84 2.14 35 0.36 0.35 0.27 0.35 3.13 3.00 2.55 3.01 0.21 0.18 0.22 2.05 1.84 2.00 40 0.40 0.39 0.31 0.39 3.02 2.91 2.55 2.91 0.22 0.20 0.23 1.92 1.83 1.88 45 0.45 0.44 0.35 0.44 2.92 2.83 2.55 2.81 0.23 0.22 0.24 1.80 1.83 1.76 50 0.50 0.49 0.40 0.48 2.82 2.74 2.55 2.71 0.23 0.24 0.24 1.69 1.83 1.65 55 0.56 0.54 0.45 0.53 2.72 2.66 2.55 2.62 0.24 0.26 0.24 1.58 1.83 1.54 60 0.62 0.60 0.52 0.59 2.63 2.58 2.55 2.53 0.24 0.28 0.25 1.48 1.82 1.45 65 0.68 0.66 0.59 0.64 2.54 2.51 2.55 2.44 0.24 0.31 0.25 1.39 1.82 1.36 70 0.74 0.73 0.67 0.70 2.45 2.43 2.55 2.36 0.24 0.34 0.25 1.30 1.82 1.27 75 0.81 0.80 0.76 0.76 2.37 2.36 2.55 2.28 0.24 0.37 0.25 1.22 1.82 1.19 80 0.88 0.87 0.86 0.82 2.29 2.29 2.55 2.20 0.24 0.41 0.25 1.14 1.82 1.12 85 0.95 0.94 0.88 2.21 2.22 2.12 0.24 0.24 1.07 1.05 90 1.02 1.02 0.94 2.13 2.16 2.05 95 1.09 1.10 1.01 2.06 2.09 1.98 100 1.17 1.19 1.07 1.99 2.03 1.91 105 1.24 1.27 1.14 1.92 1.97 1.85 110 1.32 1.20 1.85 1.78 115 1.39 1.27 1.79 1.72 120 1.47 1.33 1.73 1.66 125 1.54 1.40 1.67 1.61 130 1.62 1.46 1.61 1.55 McGill, and Laval Universities who assisted with the field work. Special thanks are merited by L. Natanson for her contribution to the organization and devel- opment of the sampling progi-am, to T. Miller for sup- plying measurements of larval fish standard length based on video recordings at sea, G. Chaumillon, L. Fortier, and J. Himmelman for comments on the manuscript, J. Butler for allowing access to his im- age analyzer, and to two anonymous reviewers who critically reviewed the manuscript and offered many helpful suggestions. L. Michaud and M.-A. Remillard assisted in the laboratory. The officers and crew of the RV Petrel V provided excellent assistance during the execution of the field work. Literature cited Andersen, T., and E. Moksness. 1988. Manual for reading daily increments by the use of a computer programme. Flodevigen Meldinger NR. 4, 37 p. Baker, A. C, M. R. Clarke, and M. J. Harris. 1973. The N.I.O. combination net (RMT 1+8) and further developments of rectangular midwater trawls J. Mar. Biol. Assoc. U.K. 53:167-184 Bolz, G. R„ and R. G. Lough. 1983. Growth of larval Atlantic cod, Gadus mohrua, and haddock Melanogrammus aeglefinus, on Georges Bank, spnng 1981. Fish. Bull. 81:827-836. 1988. Growth through the first six months of Atlantic cod, Gadus mohrua. and haddock Melanogrammus aeglefinus, based on daily otolith increments. Fish. Bull. 86:223-235. Brander, K., and P. C. F. Hurley. 1992. Distribution of early-stage Atlantic cod iGadus morhua ), haddockiMelanogrammus aeglefinus), and witch flounder iGlyptoeephalus cynogtossus) eggs on the Scotian Shelf a reappraisal of evidence on the coupling of cod spawning and plankton production. Can. J. Fish. Aquat. Sci. 49:238-251. Brothers, E. B., and W. N. McFarland. 1981. Correlation between otolith microstructure, growth, and life history transitions in newly recruited French grunts [Haemulon flavolmeatum (Demarest), Haemu- lidaej. In R. Lasker and K Sherman (eds.). The early life history of fish. p. 369-374. Rapp. R.-V. Reun. Cons. Int. Explor Mer 178. 318 Fishery Bulletin 97(2), 1999 Brothers, E. B., C. P. Mathews, and R. Lasker. 1976. Daily growth increments in otoliths from lai-val and adult fishes. Fish. Bull. 74( 1 ):l-8. Campana, S. E. 1989. Otolith microstructure of three larval gadids in the Gulf of Maine, with inferences on early life history. Can. J. Zool. 67:1401-1410. Campana, S. E., and P. C. F. Hurley. 1989. An age-and-temperature-mediated growth model for cod (Gadus morhua) and haddock {Melanogrammus aeglefinus) larvae on the Scotian Shelf. Can. J. Fish. Aquat. Sci. 46:603-613. Campana, S. E., and J. D. Neilson. 1985. Microstructure of fish otoliths. Can. J. Fish. Aquat. Sci. 42:1014-1032. Drinkwater, K., B. Petrie, and W. H. Sutcliffe Jr. 1979. Seasonal geostrophic volume transports along the Scotian Shelf Estuarine Coastal Mar Sci. 9:17-27. Fahay, M. P. 1983. Guide to the early stages of marine fishes occurring in the western North Atlantic Ocean. Cape Hatteras to the southern Scotian Shelf J. Northwest Atl. Fish. Sci. 4: 1-423. Fridgeirsson, E. 1978. Embrionic development of five species of gadoid fishes in Icelandic waters. Rit Fiskideildar V(6):68 p. Geffen, A. J. 1987. Methods of validating daily increment deposition in otoliths of larval fish. In R. C. Summerfelt and G. E. Hall (eds.), The age and growth offish, p. 223-240. Iowa State Univ. Press, Ames. lA. Griffin, D., and C. Anderson. 1993. OPEN REPORT 1993/4: meteorological, hydro- graphic, and acoustic Doppler current profiler (ADCP) ob- servations on the Scotian Shelf Department of Ocean- ography, Dalhousie Univ., Halifax, Nova Scotia, Canada, 182 p. Houde, E. D. 1987. Fish early life dynamics and recruitment variability. Am. Fish. Soc. Symp. 2:17-29. Jones, C. 1992. Development and application of the otolith increment technique. In D. K. Stevenson and S. E. Campana (eds.). Otolith microstructure examination and analysis, p. 1- 11. Can. Spec. Publ. Fish. Aquat. Sci. 1 17. Koeller, P. A., P. C. F. Hurley, P. Perley, and J. D. Neilson. 1986. Juvenile fish surveys on the Scotian Shelf implica- tions for year-class size assessments. J. Cons. Int. Explor Mer 43:.59-76. Lasker, R. 1975. Field criteria for survival of anchovy larvae: the re- lation between inshore chlorophyll maximum layers and successful first feeding. Fish. Bull. 73(31:4.53-462. Laurence, G. C. 1974. Growth and survival of haddock iMelanogrammus aeglefinus) larvae in relation to planktonic prey concen- tration. .J. Fush. Res. Board Can. 31:141,5-1419. 1978. Comparative growth, respiration and delayed feed- ing abilities of lar\'al cod iGadus morhua) and haddock (Melanogrammus aeglefinus) an influenced by temperature during laboratory studies. Mar. Biol. 50:1-7. Laurence, G. C, and C. A. Rogers. 1976. Effects of temperature and salinity on comparative embryo development and mortality of Atlantic cod iGadus morhua) and haddock ) Melanogrammus aeglefinus). J. Cons. Int. Explor Mer .36; 220-228. Lough, R. G., M. Pennington, G. R. Bolz, and A. A. Rosenberg. 1982. Age and growth of larval atlantic herring, Clupea harengus L., in the Gulf of Maine-George Bank region based on otolith growth increments. Fish. Bull. 80(2):187-199. Mahon, R., and J. D. Neilson. 1987. Diet changes in Scotian Shelf haddock during the pelagic and demersal phases of the first year of life. Mar Ecol. Prog Ser 37:123-130. Moksness, E., and V. Wespestad. 1989. Ageing and backcalculating growth rate of Pacific herring (Clupea harengus pallast ) larvae by reading daily otolith increments. Fish. Bull. 87:509-518. Nishimura, A., and J. Yamada. 1984. Age and growth and larval and juvenile walleye poUockr/jeragra chalcogramma (Pallas), as determined by otolith daily growth increments. J. Exp. Mar Biol. Ecol. 82:191-205. O'Boyle, R. N., M. Sinclair, R. J. Conover, K. H. Mann, and A C. Kohler. 1984. Temporal and spatial distribution of ichthyoplankton communities of the scotian shelf in relation to biological, hydrological, and physiological features. Rapp. P.-V. Reun. Cons. Int. Explor Mer 183:27-40. Palomera, L, B. Morales-Nin, and J. Lleonard. 1988. Lai-val growth of anchovy, Engrautis encrasicolus, in the western Mediterranean Sea. Mar Biol. 99:283-291. Pannella, G. 1971. Fish otoliths: daily growth layers and periodical patterns. Science (Wash., D.C.) 173:1124-1127. Radtke, R. L. 1989. Larval fish age, growth, and body shrinkage infor- mation available from otoliths. Can. J. Fish. Aquat. Sci. 46:1884-1894. Radtke, R. L., and J. M. Dean. 1982. Increment formation in the otoliths of embryos, lar- vae and juveniles of the mummichog Fundulus hetero- clitus. Fish. Bull. 80:201-215. Radtke, R. L., and K. G. Waiwood. 1980. Otolith formation and body shrinkage due to fixa- tion in larval cod iGadus morhua). Can. Tech. Rep. Fish. Aquat. Sci. No. 929. iii + 10 p. Ricker, W. R. 1979. Growth rates and models. In W. S. Hoar, D. J. Randall, and J. R. Brett (eds.) Fish physiology, p. 677- 743. Acad. Press, New York, NY. Rothschild, B. J., and C. Rooth. 1982. Fish ecology III A foundation for REX-recruiting experiment. Univ. Miami Technical Report 82008, Miami, FL, .389 p Scott, W. B., and M. G. Scott. 1988. Atlantic fishes of Canada. Can. Bull. Fish. Aquat. Sci. 219, 731 p. Shepherd, J. G., and D. H. Gushing. 1980. A mechanism for density-dependent survival of lar- val fish as the basis of a stock-recruitment relationship. J. Cons. Int. Explor Mer 39:160-167. Simard, P., Castonguay, D. D. D'Amours, and P. Magnan. 1992. Growth comparision between juvenile Atlantic mack- erel [Scomber scombrus) from the two spawning groups of the Northwest Atlantic. Can. J. Fish. Aquat. Sci. 49:2242- 2248. Struhsaker, P., and J. H. Uchiyama. 1976. Age and growth of the nehu, Stolephorus purpureus, from the Hawaiian Islands as indicated by daily growth increments of sagittae. Fish. Bull. 74:9-17. Quihonez-Velazquez; Age validation and growth of Melanogrammus aeglefinus and Pollachius virens 319 Sutcliffe, W. H., Jr., R. H. Loucks, and K. F. Drinkwater. 1976. Coastal circulation and physical oceanography of the Scotian Shelf and Gulf of Maine. J. Fish. Res. Board Can. 33:96-115. Taubert, B. D., and D. W. Coble. 1977. Daily rings in otoliths of three species of Lepomis and Tilapia mossambica. J. Fish. Res. Board Can. 34:332-340. Ware, D. H. 1975. Relation between egg size, growth, and natural mor- tality of larval fish. J. Fish. Res. Board Can. 32:2503- 2512. Watanabe, N., K. Tanaka, J. Yamada, and J. Dean. 1982. Scanning electron microscope observations of the or- ganic matrix in the otolith of the teleost fish Fundulus heteroclitus (Linnaeus) and Tilapia nilotica (Linnaeus). J. Exp. Mar. Biol. Ecol. 58:127-134. Watanabe, Y., J. L. Butler, and T. Mori. 1988. Growth of Pacific saury. Cololabis saira, in the north- eastern and northwestern Pacific Ocean. Fish. Bull. 86:489-498. Wilkinson, L. 1990. SYSTAT; the system for statistics. SYSTAT, Inc., Evanston, IL, 677 p. Zweifel J. R., and R. Lasker. 1976. Prehatch and posthatch growth of fishes: a general model. Fish. Bull. 74:609-621. 320 Abstract.— We used hydroacoustic techniques to obtain biomass estimates ofyellowtail rockfish, Sebastes flavidus, in a lOO-km-'area off the west coast of British Columbia, Canada. The purpose of our study was to estimate sampHng variance and explore the effect of diel aggregation behavior on the precision of biomass estimates. A set of eight transects was sampled eight times: four at night, four during daylight. Although we observed a pronounced diel behav- ioral pattern of diurnal aggregation and nocturnal dispersion, we found no sig- nificant differences between nocturnal and diurnal estimates of mean biomass. Diurnal estimates showed a tendency towards higher variance, but the dif- ferences were not significant in most comparisons and were too small to in- fluence survey design. The coefficient of variation of the eight observations for any individual transect ranged from 13 to V28'7r. The coefficient of variation in biomass for the whole survey area, based on repeating the set of eight transects eight times, was IS.B*?? and the estimate of mean biomass for the survey area was 1152 t. The observed diel behavioral patterns did not, in this study, produce different estimates of yellowtail rockfish biomass. Survey time might therefore be optimized with- out concern for this source of variance for this species. Diel vertical migration by yellowtail rockfish, Sebastes flavidus, and its impact on acoustic biomass estimation Richard D. Stanley Robert Kieser Bruce M. Leaman Ken D. Cooke Biological Sciences Branch Fisheries and Oceans Pacific Biological Station Nanaimo, British Columbia, V9R 5K6 Canada E-mail address (for R D Stanley) stanieyrffidfo-mpo gc ca Manuscript accepted 2.') June 1998. Fi.sh. Bull. 97.320-3.31 (1999). Yellowtail rockfish, Sebastes flavidus, are an important component of the British Columbia (B,C,) trawl fish- ery. Annual catches of 4000-5000 t represent about 207r of the total rockfish {Sebastes spp.) landings and over 57( of the total domestic trawl landings from the B.C. coast (Rutherford, 1996). The stock as- sessments for these species have typically relied upon population dynamics models tuned with catch- per-unit-of-effort indices (Stanley, 1993) or swept-area surveys (Lea- man and Stanley, 1993). However, problems with both these methods have led us to investigate more di- rect methods. Previous hydroacoustic work in- dicated that some of the trawl- caught rockfishes are aggregating species and that they are usually found associated with specific bathymetric features (Wilkins, 1986; Leaman et al., 1990; Kieser et al., 1992; Richards et al., 1991). Submersible observations and acoustic tagging of a small number offish confirmed this association for yellowtail rockfish (Pearcy et al., 1989), although the acoustic tagging did not indicate consistent diel pat- terns (Pearcy, 1992). This aggregat- ing behavior off the bottom in pre- dictable locations suggests that this species may be a suitable candidate for hydroacoustic biomass estima- tion. However, because individual yellowtail rockfish stocks are thought to occupy large areas (Stanley, 1993; Stanley et al., 1994), such compre- hensive hydroacoustic surveys would be costly and time-consum- ing. This accentuated the need for a prior study of fish behavior and sampling variance to assess the fea- sibility of such an approach and to optimize survey effort. We were also concerned about the impact of diel aggregating behavior on the preci- sion and bias of biomass estimates (Olsen, 1990; Appenzeller and Leg- gett, 1992; Simmonds et al., 1992). Within the same genus, Richards et al. (1991) reported strong diurnal aggregation by Pacific ocean perch (S. alutus), and Wilkins ( 1986) has reported strong nocturnal aggrega- tion by widow rockfish (S. entomelas). We investigated these issues with an experimental survey in Novem- ber 1991 in waters 10 km west of Vancouver Island, British Colum- bia, Canada (Fig. 1). The objective was to estimate yellowtail rockfish abundance within a small area and, in doing so, to examine the variance within and among transects. In ad- dition, we wished to characterize diel patterns of behavior and docu- ment the effects of such behavior on the precision of the biomass esti- Stanley et al.: Diel vertical migration by Sebastes flavidus 321 ■5I°00' BRITISH \ COLUMBIA [\ -N- 50 100 -48''00'-o O en CM STUDY AREA VANCOUVER ftNAQA . Tj.S^A. KILOMETRES Figure 1 Location of study area off the southwest coast of Vancouver Island, British Co- lumbia, Canada. mates. In this report, we summarize our observa- tions from this study and discuss the implications of the results to hydroacoustic survey design for yel- lowtail rockfish. Methods Study area We first surveyed a 250-km^ area along the shelf edge off Nootka Island to locate concentrations of rock- fish (Fig. 1). This area produces 1000-3000 t of yel- lowtail rockfish per year (Rutherford, 1996). An ini- tial grid included nine parallel transects 9-13 km in length and 2.8 km apart. They were oriented south- west to northeast, perpendicular to the isobaths. The southwestern end of the track lines extended well beyond the shelf break, whereas the shallower north- eastern end extended onto the shelf to the 130-m isobath. This A series was covered during one 24-h period. The northern half of the A series showed the greatest concentrations of rockfish. From this north- ern area, we selected a B series of transects to con- duct the study (Fig. 2). No further use was made of A-series measurements because the respective transects were not comparable with the B-series transects. The distance between transects was halved to 1.4 km, and the length of each was reduced to span depths from the 140-m isobath to just beyond the shelf break (180 m). The A series had shown that essentially all fish concentrations were between these limits. The B series was repeated in the same se- quence (B9-B2) through four 24-h cycles to produce a balanced design of four diurnal and four nocturnal observations per transect. In addition to the acous- tic estimates for the eight replicates of B2-B9, we selected B5 and B6 for continuous soundings, result- ing in ten additional observations for these two transects. The bottom topography of transects B2-B9, with the exception of B8, was characterized by one or two underwater cliffs They appeared to be 10-15 m high and followed a north-south axis through the study area, landward of the shelf break (Figs. 2-4). The transects were oriented perpendicular to the axes of the cliffs. At the southernmost transect ( B9 ), the cliffs were approximately 1.5 km inshore of the shelf break. This distance increased to about 7.0 km by transect B2. The cliffs were found at a progressively shallower depth. This topographic feature is well known to fish- ermen as the "clay bumps" on the Nootka fishing grounds. Fishermen report that catches of yellow- tail rockfish can be made year-round by towing bot- tom trawl nets in an east to west (shallow to deep) direction over the cliffs. Transect B5 corresponds to a commonly used commercial fishery tow (Figs. 2-4). We excluded the results of transect Bl from our analyses. It passed over an extended underwater ridge such that the modal depth was less than 130 m rather than the targeted depth of 150 m. These depths are shallower than the November habitat 322 Fishery Bulletin 97(2), 1999 Figure 2 Area covered by the B-series transects and location of the shelf edge ( 200 ml and the cliffs in relation to the study area. range for yellowtail rockfish (Nagtegaal, 1983). We also had to exclude the third of the four nocturnal replicates of transect B4. The vessel had to slow and change course to avoid traffic, leading to an unus- able acoustic estimate. We used an average from the other three nocturnal estimates in our principal analyses to maintain a balanced ANOVA design. We also substituted a range of values in place of the miss- ing value to examine the stability of the results. Constant transects allowed us to compare behav- ior over the same location, but in doing so we sacri- ficed the improved statistical power we would have achieved with a stratified random design (Jolly and Hampton, 1991). Time of the transect was classified according to the moment the vessel passed over the cliffs, the ob- served center of fish abundance. These times were then classified as diurnal or nocturnal according to the time of sunrise and sunset for Tofino, approxi- mately 100 km southeast of the study area (Atmo- spheric Environment Service, Environment Canada). Transect length We derived two density estimates, based on a short and a long section from each completed transect, to examine the impacts of the choice of endpoints. For both sections and for each transect, we chose a stan- dard beginning and end in relation to the cliffs (Fig. 3). The ends for the short sections of the transects were defined as the shoreward and seaward limits of the nocturnal dispersion of the cliff aggregations, judged visually from the echograms for all replicates. An individual transect required about one hour to complete. For these sections, the seaward end was inshore of the shelf break. The long sections used the same shoreward or shallow end, whereas the seaward or deep end was extended to a 180-m bot- tom depth. This depth incorporated the apparent deep-water limit of the shelf break aggregations (Fig. 3). The echograms for replicate transects (Fig. 3) differ in length because of differences in vessel speed over the ground. We calculated an average surface density (g/m'^) based on the constant length for each replicate of the short and long sections from all transects. Hydroacoustic equipment and echo processing Echo integration was used to estimate fish abun- dance in the survey area (Forbes and Nakken, 1972; Clay and Medwin, 1977; Foote, 1987; MacLennan and Simmonds, 1991). The calibrated echo integration system was operated from the Canadian Coast Guard Ship, W. E. Richer. The acoustic processing equip- ment consisted of a BioSonics 38 kHz Model 101 echo sounder, BioSonics Model 111 chart recorder, BioSonics Model 121 digital echo integrator, PCM/ VCR tape recording system, and auxiliary equip- ment. The transducer components included a towed body with a Simrad ceramic transducer and armored tow cable. Stanley et al.: Diel vertical migration by Sebastes flavtdus 323 The echo integrator was programmed to ana- lyze the return echoes for a series of depth strata (range slices) from the transducer to the bot- tom (Kieser et al., 1987). Bottom tracking was obtained with a 5-m bottom buffer An echo in- tegration sequence was completed every 60 pings (1 minute), and the measured echo inten- sities were stored on a personal computer. Echo integrator and chart recorder thresholds were set to 0.2 V; time-varied gain (TVG) was set to 20 log R. Thus all integrated echoes were dis- played on the echogram. At this threshold level, noise was negligible at all depths of interest. Vessel speed was approximately nine knots. The integrator output was processed with cus- tom software to exclude extraneous signals from the ocean bottom, as well as noise and echoes from other unwanted sources in the water column. A target strength (TS) of -32 dB/kg was used to convert the measured backscatter cross sec- tion to fish volume density (g/m'^) and fish sur- face density tg/m'") estimates. This value was obtained from a review of the literature (Foote, 1987; Kieser, 1992); no in situ rockfish TS were available at that time. Only a relative TS was re- quired because the study focused on the acoustic availability of rockfish and relative rather than absolute biomass and variance estimates. Biomass estimates were obtained by extrapo- lating surface densities to cells that were bounded by lines equidistant between adjacent transects and by lines perpendicular to the transect and passing midway between measure- ments. An estimate of the total fish biomass for the area was obtained by summing the prod- ucts of surface densities and cell areas. B5-D 1401 B5-N 2158 B6-D 1302 B6-N 2059 B7-D 1155 B7-N 2002 Figure 3 Diurnal (D) and nocturnal (N) echograms from transects B.5, B6. and B7 for November 17. "Sh" identifies the shallow end of both long and short transects. "S" and "L" identify the deep ends of the short and long sections of the transects. Time shown repre- sents the approximate time the vessel passed over the cliffs. L S Fishing and sampling We conducted three bottom trawl tows (Table 1 ). Two were conducted along the path of transect B5 and the other along transect B6. All three targeted diur- nal aggregations at the cliffs. The tows ran from shal- low to deep and were terminated at the seaward side of the cliffs, prior to encountering the aggregations near the shelf break. Planned additional bottom and midwater trawl fishing was curtailed owing to opera- tional difficulties aboard the vessel. Analysis of variance Transect series 1, 2, 3, and 4 were conducted 17-18, 20-21, 22, and 24-25 November (Table 2). We ini- ' tially tested for diel, transect, and series effects with a three-factor ANOVA: where D ,jk I-' D,jf, =iJ + a,+lij+ 7^, + f , ijk' = surface density for transect /, day or nighty, and observation (or series) k; = the overall mean; = the transect effect; = the diel effect. = the influence of the date (series); and /, = the error term, normally distributed with a mean of 0 and variance of a'^. To test whether diel variances were heterogeneous, we conducted a one-way randomization ANOVA lim- iting the factors to the diurnal-nocturnal effect. We conducted this test on a scaled version of the obser- vations. Because it was obvious that the variance 324 Fishery Bulletin 97(2), 1999 among transects was, in part, proportional to the mean, we removed the transect effects by scahng all the observations. Each observation was multiplied by the overall mean for the data set (n=64) divided by the mean for all eight observations (diurnal and nocturnal) of the transect. Thus the variation in the data set was reduced to the scaled "within transect" and diel effects. Following examination of the balanced treatment of all transects, we examined the observations for transects B5 and B6 (Table 3). These data included observations from the four series as well as additional nonpaired observations. The two transects were ex- B5-0603 B5-()743 65-0943 B5-1153 B5-I437 B5- 1 607 B5-1817 Figure 4 Echogram.'; oftran.scct B.") over 12 hours .showing dawn and du.sk transition of fi.sh distribution over the cliffs. Time shown represents the approximate time the vessel passed over the cliffs. amined separately. Homogeneity of variance between diel periods was examined by using an F-ratio test. We used nonparametric randomization tests ( Manly, 1991 ) to examine the sources of variance owing to con- cern about the small sample sizes, non-normality, and heterogeneity of variance. In particular, we hypoth- esized that variance among repeated transects would be proportional to the biomass estimates and, owing to the extreme densities and localization of the diurnal aggregations, that diurnal variance would be signifi- cantly gi'eater than nocturnal variance. Randomization involves resampling without re- placement from one parent population. The observed value of the response statistic (for example, mean diurnal biomass minus mean nocturnal biomass) is compared with a large number of responses gener- ated by treating all obsei'\'ations (diurnal and noc- turnal) as coming from one parent population. Each of 4999 resamplings randomly allocates the obser- vations into two groups, each time simulating new diurnal and nocturnal sets. The observed difference in mean biomass is then compared with the 4999 simulated differences. If the observed difference in biomass is significant (a<0. 05), then the observed dif- ference should be greater than 95^^ of all the simu- lated differences. Results Fish behavior and species composition Four aggregation types were observed on the echo- grams. Each was associated with a different habi- tat. The first, our targeted group, was located near the cliffs over depths of 150-160 m (Fig. 3). These dense schools were located near the bottom during each day. then the fish dispersed vertically and hori- zontally at twilight. By early nighttime, the scattered targets were distributed from 80 to 150 m. The diel cycle was completed by downward migration and rapid reformation of schools at dawn. With the exception of benthic species like fiatfish, which produce a negligible contribution to the acous- tic measurement, the species composition in the catch from this aggregation type was over SS'^^f yellowtail rockfish in the three tows (Table 1), indicating that this species dominated in the diurnal aggregations targeted near the cliffs. The continuous transition from day to night distributions (Fig. 4) implies that the nocturnal aggregations were also yellowtail rock- fish, but it was not possible to confirm this assump- tion during the cruise. Subsequent discussions with fishermen familiar with nighttime trawling in the area supported the assumption that the nocturnal I Stanley et al,: Diel vertical migration by Sebastes flavidus 325 Table 1 Catch composition (%) by weight and by species from diurnal bottom trawl tows ("tr" signifi BS trace, <0.05%). Tow Total 1 2 3 Rockfish Silvergray rockfish [Sebastes brevispinis) 3.8 4.6 7.7 4.0 Widow rockfish (S. entomelas) 0.9 0.5 1.3 0.8 Yellowtail rockfish (S. flavidus) 88.2 83.2 38.5 85.9 Bocaccio (S. paiicispiriis) 7.1 2.8 tr 6.2 Other rockfish (S. spp.) tr tr 3.8 0.2 Other species (with swimbladders) tr 0.3 2.6 0.1 Other species (without swimbladders) tr 8.4 46.2 2.8 Total catch ( kg 1 1989 393 78 2460 Table 2 Biomass estimates (t) for the short and long sections of the transects ( 'na" indicates not available, measurement aborted owing to vessel traffic). Transect Series ] 2 3 4 Short Long Short Long Short Long Short Long Day 2 93 119 65 70 96 104 170 188 3 162 162 228 232 114 118 103 106 4 98 99 37 50 126 134 46 60 5 124 130 84 92 124 136 92 124 6 424 448 520 522 32 32 396 402 7 148 148 39 40 186 194 111 116 8 4 4 58 58 161 208 18 20 9 187 187 209 278 296 345 118 123 Night 2 181 200 112 124 116 123 59 68 3 92 113 152 166 85 90 124 134 4 56 112 77 103 na na 74 84 5 128 230 112 162 128 180 92 100 6 156 216 150 180 476 512 128 144 7 198 212 114 138 224 235 229 241 8 185 224 142 169 131 173 146 157 9 76 126 178 202 240 357 168 208 targets would probably have been yellowtail rock- fish. Commercial fishery records fi-om these fishing grounds consistently indicate a predominance of yel- lowtail rockfish in midwater and bottom trawl tows. The daytime tows support the assumption that the aggregation over the cliff was that of yellowtail rock- fish, and the nocturnal dispersion of these aggrega- tions is obvious from the night transects. These ag- gregations appear almost as dense as the bottom on our daytime echograms. The cliffs can be identified only in the echogram during the night (Figs. 3 and 4). We are not, however, suggesting that all fish in the transect were yellowtail rockfish, nor are we suggest- ing that the diurnal aggregations at the cliff repre- sented all the yellowtail rockfish that were present in the transects. The nocturnal "cloud" may have included fish that were close to the bottom but not adjacent to the cliffs. Figure B6-N indicates that there was some move- ment from off the edge onto the long and short ver- 326 Fishery Bulletin 97(2), 1999 Table 3 Biomass estimates (t) for the short and long sections of B5 and B6 (all observations . Type Transect Series — 1 2 3 4 — — — — — — Day Short 5 124 84 124 92 64 182 116 610 38 Long 5 130 92 136 100 172 182 132 632 332 Short 6 424 520 32 396 300 500 312 92 Long 6 448 552 32 402 308 508 320 124 Night Short 5 136 128 112 128 92 28 170 120 102 132 Long 5 188 230 162 180 124 28 212 156 140 288 Short 6 170 156 150 476 128 238 212 240 176 294 244 Long 6 192 216 180 512 144 272 256 262 232 244 280 sions of the transect, suggesting some added contami- nation; however most of the backscatter was still con- centrated near the cliffs. A second aggregation type was represented by the schools observed at the shelf break, 10-20 m deeper than the base of the cliffs (Fig. 3). These schools, al- though exhibiting diel behavior similar to that of the first group, remained closer to the bottom during daytime. The greater depth of the aggregations ( 160- 180 m), their closer proximity to the shelf edge, and their stronger bottom affinity suggested to us, as well as to various fishermen with whom we consulted, that these aggi-egations probably consisted of a combina- tion of Pacific ocean perch (S. alutiis). canary (S. pinniger), redstripe (S. proriger), sharpchin (S. zacen- trus). and silvergray iS. breuispinis) rockfish (Leaman and Nagtegaal, 1982, 1986; Leaman et al., 1990). The third and fourth target types included small schools or individual targets within the shallow end of the transects and scattered distributions over much deeper water beyond our transects and the shelf edge. The shallower signals were thought to be plankton or small fishes. Fishermen reported that attempts to fish on these aggregations were unsuccessful and suggested that the source of the signals was too small to be re- tained by their gear. No in situ target strength mea- surements were conducted. Signals in deeper water, off the edge of the shelf break, have been shown in other studies to be a deep plankton layer or hake (Merluccius productiis) (SaundersM. Signals close to the slope off the edge could also be generated by side echoes. Analysis of variance The randomization tests provide the percent of the random re-orderings that exceed the value of the Table 4 Results of three-factor randomization ANOVA (df=degrees of freedom; **>19c significance level). Percentages indicate the 'i of 4999 random combinations of the observed data which resulted in a difference between treatments greater than the observed difference. % greater than observed Factor df Short section Long section Diurnal and nocturnal 1 87.94 Transect 7 **0.08 Series 3 80.. 50 Error 53 Total 64 47.40 **0.08 66.89 ' Saunders, M. 1991. Pacific Biological Station, Nanaimo.B. C, Canada V9R 5K6. Personal conimun. observed response. Statistical significance is assumed when fewer than 5% of the re-orderings exceed the observed response. The three-factor ANOVA indicated for both short and long transect sections (Table 4) that 1 there was no significant difference in mean bio- mass between night and day; 2 there was a highly significant difference in mean biomass among transects; and 3 there was no significant difference in mean bio- mass among different series (over time). Bartlett's test for homogeneity of variance indicated no significant difference between the variance of di- urnal and nocturnal observations, although the trend was towards greater diurnal variance (Table 5). We repeated the same basic analysis using the larger but nonpaired set of observations for transects B5 and B6 (Table 3). Each transect was analyzed separately to test for significant differences between diurnal and nocturnal mean biomass and for homo- Stanley et al.: Diel vertical migration by Sebastes flavtdus 'ill Table 5 Results of Bartlett's test for significant difference between variance of diurnal and nocturnal observations (X" for P=0.01 is 2.706; x^ for P=0.05 is 3.841 ). Short section Long section 2.45 2.23 geneity of variance between the two diel periods (Table 6). These data included the observations from the comparisons above; therefore, the data sets and tests are not independent of the main ANOVA. Results for the short segments for B5 and B6 also showed no significant differences between mean noc- turnal and diurnal biomass. However, unlike the re- sults of the expanded B5 data set and the original all-transect test, diurnal variance for B6 was signifi- cantly higher than nocturnal variance for this transect. Missing observation The impact of using an average value (69 t) for the missing nocturnal observation was examined by con- ducting the randomization tests with minimum and maximum values of 56 t and 76 t for the nocturnal observations from B4. The alternate values had a negligible impact on the results. Nocturnal dispersion beyond the shallow end of the transect During one night series, the echograms indicated that the cliff aggregations dispersed beyond our empiri- cally chosen shallow endpoint in transects B5-B9. We therefore slightly underestimated the nocturnal abundance of the cliff aggregations this one night. We re-analyzed the biomass for these transects with the new endpoints and then repeated the statistical analyses for the B2-B9 and B5-B6 series only. The changes to the one night's observations did not af- fect the results significantly. Table 6 Results of comparison of expanded observation set from transects B5 and B6 ( * indicates > 5% level of significance). Percentages indicate the % of 4999 random combinations of the observed data which resulted in a difference between treatment means greater than the observed difference in means. % greater than observed Long section 22.54 15.90 10.08 1.60* Short Transect Factor section 5 Diurnal and nocturnal 35.79 F-ratio 16.08 6 Diurnal and nocturnal 5.28 F-ratio 16.10 Table 7 Coefficients of variation of biomass estimates ( B ) among transects and series (short and long sect ionsl. Area n Short section Long section B(t) CVJc B(t) CV% Total area Day 4 1167 1 7.7 1256 7.9 Night 4 1137 1 19.8 1402 19.3 Day+ 8 1152 1 13.9 1329 15.4 Night CV% range CV% range Within transects Day 4 19.8-117.8 19.0-128.5 Night 3-4 14.9-73.0 14.3-64.1 Day+ 7-8 16.6-65.8 31.4-68.1 Night Among transects Day 4 54.3-106.7 58.4-100.6 Night 4 24.5-67.3 20.6-62.2 Variance in transect and total biomass estimates Independent total biomass estimates for the study area were estimated as the sum of the transect bio- mass estimates from each series (Table 7) . The over- all average was 1152 t with a CV of 13.9*7^ among series for the short sections. For the long sections, the average was 1329 t with a CV of 15.4%. The within-transect CVs ranged from 16.6% to 65.8% for the eight transects (short sections). Aglen (1983) derived an empirical formula for pre- dicting the CV based on the ratio of linear distance surveyed (L) over the square root of the surveyed area (A), which he called the "degree of coverage" or DOC: Predicted CV = 0.5 L A -0 41 The variance among the eight series replicates of the area were significantly less than the predicted value 328 Fishery Bulletin 97(2), 1999 from Aglen's formula whereas the variance for the combinecd area of B5 and B6 was higher (Table 8). In both cases, the observed variance among surveys was typical of acoustic surveys. Discussion We have assumed that the dominant part of the acoustic signal selected from the short transects rep- resented yellowtail rockfish, a conclusion supported by fishing results and the continuity on the echo- grams between diurnal and nocturnal distribution. The longer sections of the transects, which indicated about 16% (5-30%) more total biomass, probably included other species of deeper-dwelling rockfishes. Yellowtail rockfish share with other species a diel pattern of diurnal schooling and nocturnal dispersal (Simmonds et al., 1992). We observed that the schools reform at dawn near a minor but nearly continuous topographic feature that crosses the study area on a north-south axis. The strong and persistent affilia- tion of yellowtail rockfish with this small cliff is con- sistent with long-term commercial fishing records at this location. The diurnal schools are near the bot- tom at dawn then rise slowly during the day, dis- persing at dusk. This is congruent with fishermen's observations that early morning is the most produc- tive time for bottom trawling for this species. The fidelity to topographic features is consistent with reports of Pearcy (1992) and Carlson and Haight ( 1972). Pearcy's reports on acoustically tagged speci- mens includes an example of "homing" near to the original capture sites, even as far as 2 nmi overnight. Because we conducted the study at only one time of year, we could not examine how diel behavior would vary with season. Fishermen report that yellowtail rockfish occupy the "cliffs" at all times of year. They also report that the dawn tows are the best fishing on a year-round basis. We would suggest that if diel vertical movement is triggered by or at least associ- ated with light intensity at depth, then the effect would be more obvious during summer months when light intensity increases and the twilight period shortens. It might also be more dramatic on brighter days. The diversity of species, especially near the shelf edge, illustrates that a difficult issue facing acoustic assessment of rockfish will be species identification. This process will be complicated by the difficulty of trawling much of the habitat. Our study was based on single-beam and single-frequency acoustic obser- vations. Encouraging results on species identifica- tion have been obtained by more advanced acoustic analysis of individual fish schools (Kieser and Langford, 1991; Scalabrin and Masse, 1993), com- parison of day and night survey data ( Gerlotto, 1993 ), and multifrequency and wide band observations (Simmonds and Armstrong, 1990; Zakharia, 1990). In this context, innovative transducer design and deployment will be important to obtain more detailed acoustic data from individual fish and schools near the bottom.'^ Supplemental information on habitat and depth preference by species, perhaps developed from simultaneous use of submersible devices or side- scan bathymetry, may help to estimate species' pro- portions (Richards et al., 1991). Diurnal versus nocturnal density and biomass estimates Although the echograms showed differences in diel distribution for yellowtail rockfish, they did not in- dicate the extreme densities which lead to acoustic shadowing (Foote, 1990). In contrast to other stud- ies (summarized in Appenzeller and Leggett, 1992), which report higher density estimates from night obsei-vations, our day and night estimates were simi- lar. We conclude that, for yellowtail rockfish, diel be- havior patterns do not bias hydroacoustic biomass estimates. It remains possible, however, that the similarity between diel periods is purely fortuitous in that the various factors that could affect estima- tion, such as movement in and out of the study area - Dalen, J., and H. Bodholt. 1991. Deep towed vehicle for fish abundance estimation, concept and testing. ICES Council Meeting (CM) 1991/B:53. 13 p. [Mimeo.l Table 8 Observed coefficient of variation (CV) of biomass estimates from short section in comparison with predicted CV from formula by Aglen (1983). Sample size, n. is the number of replicate coverages of either the area represented by transects B5 and B6, or the overall area of B2-B9. Transect Distance Area Predicted ;; Section ' (km) (km-) DOC C\'^i Observed Mean estimated CV7c biomass (t) B5-6 B2-9 19 Short 11.0 15.2 2.82 32.7 8 Short 39.6 55.5 5.33 25.2 48.0 397 13.9 1152 Stanley et al.: Diel vertical migration by Sebastes flavtdus 329 or diel variability in tilt angle, may simply have can- celled each other out. We assumed that there are no diel changes in fish target strength. Such changes have been observed for other species and have been linked to swimming behavior or mean tilt angle (Miyashita et al., 1995; Buerkle and Sreedharan, 1981; Olsen, 1990; Misund, 1997). Target strength variation between night and day may be smaller for yellowtail rockfish because these fish tend to remain in the same general depths. They are not moving from the bottom to surface wa- ters and therefore not encountering large relative changes in pressure. The observed diel behavior of increased aggrega- tion during the day led us to hypothesize that vari- ance in biomass should be greater for diurnal obser- vations. We observed that yellowtail rockfish move towards the cliffs at dawn. We assumed that they would concentrate further along that narrow band to produce discrete schools. We viewed the noctur- nal distribution as a relatively dispersed band (three- dimensional cloud) of individuals in the general vi- cinity of the cliffs. We assumed that the diurnal dis- tribution, which lay along the orientation of the cliff, would be a much narrower string of individual schools and hence of highly varying density, perhaps to the extent of approximating a "beaded pattern" of distinct schools following the length of the cliff We expected that diurnal estimates derived from transects that pass perpendicular to the cliffs would be highly variable because the path of the acoustic beam could range from "missing entirely" to "com- pletely ensonifying" a dense school. We found that, for the short transect version of the all-transect data, the variance of diurnal biomass estimates was not significantly greater than that of nocturnal biomass estimates. Diurnal variance was significantly higher in the expanded set of the B5- B6 set of observations but only for the long version of transect B6. The weak or mixed indication of higher diurnal variance implies that the apparent aggregation to- wards the cliffs, evident in the echograms, is not matched to the same degree by aggregation along the axis of the cliffs. We expected discrete diurnal schools following the axis of the cliffs. This tendency to aggi-egate into a continuous band offish along the cliffs, as opposed to discrete schools, may be unique to this area where there is a longitudinal topographic feature along the preferred depth. In areas of the coast lacking in such a linear feature, fish may tend to aggregate over a specific point, such as a pinnacle, producing greater daytime variance among estimates of biomass. Results of this experiment, however, do not support the hypothesis that estimates from the dispersed condition will show less variance for yel- lowtail rockfish. The similarity in variance between nocturnal and diurnal periods indicates that no sub- stantial gains in efficiency or precision can be achieved by sampling during one or other of the diel periods. The among-transect variance overwhelms the other sources of variance even within a small coastal area. From the perspective of two-stage sampling (among-transect or within-transect variance), preci- sion of the overall estimator is reduced by allocating sampling effort in proportion to variance contributed by each stage. Because among-transect variance is much greater, survey design should maximize the number of transects at the expense of replicating transects. The only exception to this principle would arise when the cost of replicate samples is much lower. Because the cost of collecting a replicate transect esti- mate is almost equal to that for an additional transect over the scale we are considering, it would be more ef- ficient to maximize the number of different transects. By repeating the survey eight times, we were able to determine the variance of the biomass estimate directly rather than by inferring it indirectly through geospatial analysis of "within" variance (Petitgas, 1993 ). This calculation of "among" survey variance fol- lows fi-om work by Williamson ( 1982), who resampled from the individual samples averaged over one minute within a transect, and by Robotham and Cas- tillo (1987), who bootstrapped cumulative transect observations as we did. There is no question that a formal investigation of the spatial impact on vari- ance would require treatment at the granular level of the observations, which the geospatial methods provide. These procedures also facilitate an investi- gation of the impact of additional explanatory vari- ables and development of model-based inference. However, the requirements of our study were real- ized by the simpler transect-based analysis. Total biomass estimate and survey design The overall biomass estimate of 1152 t for the study area appears consistent with overall coastal bio- mass estimates of 50,000-60,000 t for a stock that is assumed to extend from the study area to the border of B.C. and Alaska (Stanley 1993). The fact that the overall coefficient of variation among the eight se- ries for the study area was under 14% is encourag- ing and better than that predicted from the formula of Aglen (1983). This precision may not, however, apply to all rockfish species. Wilkins (1986) found that the CV for widow rockfish was 2-3 times higher in spite of a more comprehensive estimation proce- dure that included the use of side-scan sonar in con- junction with echosounding. We can also expect 330 Fishery Bulletin 97(2), 1999 |»* I ill greater variation if repeated estimates were con- ducted at greater time intervals. Over a one-week period and an area of over 100 km^, we have demonstrated precision (CV=15%) in acoustic biomass estimates that is acceptable in a stock assessment context for yellowtail rockfish. Vari- ance among replicated transects is low enough that survey effort can be distributed to maximize the num- ber of different transects, either to increase the area of the survey or the density of coverage. We have shown that a survey of yellowtail rockfish can be conducted throughout the diel cycle and thus reduce survey costs. This study also indicates that yellowtail rockfish can aggregate within a well-defined bathymetric range near topographic features. If these tendencies are consistent over the whole range of this species, they may provide the basis for stratification and pos- sible further gains in efficiency. Our results indicate that a simple systematic transect design with transects oriented perpendicular to the long axis of the fish con- centrations (the edge of the continental shelf) is a sat- isfactory choice for the elementary sampling distance unit (ESDU) (Simmonds et al., 1992; Simmonds and Fryer, 1996). If the preferable depth range is narrow over most of the coastline and the survey must thus cover a long nan-ow corridor, then a systematic zig-zag would be preferable (Simmonds et al., 1992). The affiliation of yellowtail rockfish with a minor topographic feature within a depth range indicates that the density distribution of yellowtail rockfish is "nonstationary," in that the densities will not be ran- domly distributed in a study area or stratum, even after bathymetric stratification (Simmonds et al., 1992). This is an important characteristic of the spe- cies and should be noted if more advanced survey design and analysis procedures, such as geostatistical spatial averaging and cokriging, are to be investi- gated (Foote and Stefansson, 1993; Marcotte, 1991). Although the results of the experiment support the potential for acoustic estimation of this species, the generality of the conclusions will have to be tested over a larger scale and more varied habitat. The study site was chosen carefully to minimize the unknowns, in particular the presence of other species. The hypoth- eses will have to be re-examined over different depths and topography. It is also possible that the annual cycle of maturation, mating, and parturition for these live- bearing fish may be associated with different behavior. Acknowledgments We thank R. Kronlund, J. Schweigert, and three anonymous reviewers for constructive reviews of earlier drafts. Literature cited Aglen, A. 1983. Random errors of acoustic fi.sh abundance estimates in relation to the survey grid density applied. In O. Nakken and S. C. Venema (eds.). Fisheries acoustics sym- posium, Bergen, Norway, 21-24 June 1982, p. 293- 298. FAO Fish. Rep. 300, 331 p. Appenzeller, A. R., and W. C. Leggett. 1992. Bias in hydroacoustic estimates of fish abundance due to acoustic shadowing; evidence from day-night sur- veys of vertically migrating fish. Can. J. Fish. Aquat. Sci. 49:2179-2189. Buerkle, U., and A. Sreedharan. 1981. Acoustic target strengths of cod in relation to their aspect in the sound beam. II: Contributed papers, discus- sion, and discussion and comments 1981, p. 229- 247. Meeting on hydroacoustical methods for the estima- tion of marine fish populations, 25-29 June 1979. Clay, S. C, and H. Medwin. 1977. Acoustical oceanography: principles and applica- tions. John Wiley & Sons, New York, NY, 544 p. Foote, K. G. 1987. Fish target strength for use in echo integration surveys. J. Acoust. Soc. Am. 73:1932-1940. 1990. Correcting acoustic measurements of scatterer den- sity for extinction. J. Acoust. Soc. Am. 88:1543-1546. Foote, K. G., and G. Stefansson 1993. Definition of the problem of estimating fish abun- dance over an area from acoustic line-transect measure- ments of density ICES J. Mar. Sci. 50:369-381. Forbes, S. T., and O. Nakken. 1972. Manual of methods for fisheries resource survey and appraisal. Part 2: The use of acoustic instruments for fish- eries abundance estimation. FAO. Rome, 138 p. Gerlotto, F. 1993. Identification and spatial stratification of tropical fish concentrations using acoustic populations. Aquat. Living Re.sour 6(31:243-254. Jolly, G. M., and I. Hampton. 1991. Some problems in the statistical design and analysis of acoustic surveys to assess fish biomass. Rapp. F.-V. Reun. Cons. Int. Explor. Mer 189:41.5-421. Kieser, R., and G. Langford. 1991. An application of spatial analysis to fisheries acoustics. In H. Schreier, S. Brown, P. O'Reilly, and P. J. Meehan (eds.), GIS'91. Applications in a changing world; Vancouver. February 12-15, 1991. p. 33,5-339. Minister of Supply and Services Canada. Cat. No. FO 2919/153E. 442 p. Kieser, R., B. M. Leaman, P. K. Withler, and R. D. Stanley. 1992. W. E. RICKER and EASTWARD HO cruise to study the effect of trawling on rockfish behaviour. October 15- 27. 1990. Can. Man. Rep. Fish. Aquat. Sci. 2161, 84 p. Kieser R., T. J. Mulligan, M. J. Williamson, and M. O. Nelson. 1987. Intercalibration of two echo integration systems based on acoustic backscattering measurements. Can. J. Fish. Aquat. Sci. 44:562-572. Leaman, B. M., and D. A. Nagtegaal. 1982. Biomass estimation of rockfish stocks off the west coast off the Queen Charlotte Islands during 1978 and 1979. Can. MS. Rep. Fish. Aquat. Sci. 1652, 46 p. 1986. Identification of species assemblages and results of management applications for shelf and slope rockfishes off British Columbia. In Proc. Int. Rockfish Symp., Univ. Alaska Sea Grant Rep. 87-2. p. .309-328. Stanley et al,: Diel vertical migration by Sebastes flavidus 331 Leaman, B. M., R. Kieser, P. Withler, and R. D. Stanley. 1990. W. E. RICKER hydroacoustic cruise to study rock- fish behaviour off northern Vancouver Island, March 14- 23. 1990. Can. MS. Rep. Fish. Aquat. Sci. 2091, 63 p. Leaman, B. M., and R. D. Stanley. 1993. Experimental management programs for two rock- fish stocks off British Columbia, Canada. In S. J. Smith, J. L. Hunt and D. Rivard leds.l Risk evaluation and bio- logical reference points for fisheries management, p. 403- 418. Can. Spec. Publ. Fish. Aquat. Sci. 120:viii -i- 442 p. MacLennan, D. N., and E. J. Simmonds. 1991. Fisheries acoustics. Fish and Fisheries Series, Chapman and Hall, London, 336 p. Manly, B. F. J. 1991. Randomization and Monte Carlo methods in biology. Chapman and Hall, New York, NY, 281 p. Marcotte, D. 1991. Cokriging with MATLAB. Computers and Geo- sciences 17l9):126.5-1280. Miyashita, K, I. Aoki, and T. Inagaki. 1995. Orientation isada krill (Euphausia pacifica) in rela- tion to acoustical observation. Bull. Jpn. Soc. Fish. Oceanogr. 59 (31:235-240. Misund, O. A. 1997. Underwater acoustics in marine fisheries and fish- eries research. Rev. Fish. Biol. Fish. 7:1-34. Nagtegaal, D. A. 1983. Identification and description of assemblages of some commercially important rockfishes (Sebastes spp.) off Brit- ish Columbia. Can.Tech.Rep.Fish.Aquat.Sci.No. 1183.82p. Olsen, K. 1990. Fish behaviour and acoustic sampling. Rapp. P.-V. Reun. Cons. Int. Explor Mer 189:147-158. Pearcy, W. G. 1992. Movements of acoustically-tagged yellowtail rockfish Sebastes flavidus on Heceta Bank, Oregon. Fish. Bull. 90:726-735. Pearcy, W. G., D. L. Stein, M. A. Hixon, E. K. Pikitch, W. H. Barss, and R. M. Starr. 1989. Submersible observations of deep-sea reef fishes of Heceta Bank, Oregon. Fish. Bull. 87:955-965. Petitgas, P. 1993. Geostatistics for fish stock assessments: a review and an acoustic application. ICES. J. Mar Sci. 50:285-298. Richards, L. J., R. Kieser, T. J. Mulligan, and J. R. Candy. 1991. Classification offish assemblages based on echo in- tegration surveys. Can. J. Fish. Aquat. Sci. 48(7):1264- 1272. Robotham, H., and J. Castillo. 1987. The bootstrap method: an alternative for estimating confidence intervals of resources surveyed by hydroacoustic techniques. Rapp. P.-V. Reun. Cons. Int. Explor. Mer 189:421-424. Rutherford, K. L. 1996. Catch and effort statistics of the Canadian ground- fish fishery on the Pacific coast in 1993. Can. Tech. Rep. Fish. Aquat. Sci. 2097,97 p. Scalabrin, C, and J. Masse. 1993. Acoustic detection of the spatial and temporal distri- bution offish shoals in the Bay of Biscay. Aquat. Living Resources 6(3):269-283. Simmonds, E. J., and F. Armstrong. 1990. A wide band echosounder: measurements on cod, saithe, herring, and mackerel from 27 to 54 kHz. p. 381- 387. In W. A. Karp (ed.). Developments in fisheries acous- tics: a symposium held in Seattle, 22-26 June 1987. Rapp. PV. Reun. CIEM 189, 442 p. Simmonds, E. J., and R. J. Fryer, 1996. Which are better, random or systematic acoustic sur- veys? A simulation using North Sea herring as an example. ICES J. Mar Sci. 53:39-50. Simmonds, E. J., N. J. Williamson, F. Gerlotto, and A. Aglen. 1992. Acoustic survey design and analysis procedures: a comprehensive review of current practice. ICES Coop. Res. Rep. 187, 127 p. Stanley, R. D. 1993. Shelf rockfish (silvergray, yellowtail, canary, widow rockfish). In B. M. Leaman and M. Stocker(eds.), Ground- fish stock assessments for the west coast of Canada in 1992 and recommended yield options for 1993, p. 245-335. Can. MS. Tech. Aquat. Sci. 1919, 407 p. Stanley, R. D., B M. Leaman, L. Haldorson, and V. M. O'Connell. 1994. Movements of tagged adult yellowtail rockfish, Sebastes flavidus, off the west coast of North America. Fish. Bull. 92:655-663. Wilkins, M. E. 1986. Development and evaluation of methodologies for assessing and monitoring the abundance of widow rock- fish, Sebastes entomelas. Fish. Bull. 84:287-310. Williamson, N. J. 1982. Cluster sampling estimation of the variance of abun- dance estimates derived from quantitative echo sounder surveys. Can. J. Fish. Aquat. Sci., 39:229-231. Zakharia, M. E. 1990. Variations offish target strength induced by its move- ment; a wideband experiment. In W. A. Karp (ed.). De- velopments in fisheries acoustics: a symposium held in Se- attle, 22-26 .June 1987, p. 398-404. Rapp. P.V. Reun. CIEM 189, 442 p. 332 Abstract.— Disturbances to harbor seals. Phoca vitulina richardsi, during 1991 and 1992 pupping seasons were observed at Puffin Island, Clements Reef, and Skipjack Island in Washing- ton state. Harassment (> one seal en- tering the water) of seals ashore was common (>71'r of survey days) and pri- marily caused by powerboat operators approaching to observe seals. Recovery (number of seals on a haul-out site re- turned to preharassment levels ) follow- ing a harassment was less at Puffin Is- land ( 19^* ) than at Clements Reef (54'7f ) and Skipjack Island (459^). Addition- ally, seals were more vigilant (P<0.003) at Puffin Island than at the other two locations. These results indicated that seals at Puffin Island were less toler- ant of disturbance than seals at other sites. This could possibly be attributed to a greater (P<0. 05) percentage of pups ashore (IVJ) than at Clements Reef (39^ ) and Skipjack Island (3'7, ). Because of this, we expected that powerboats would disturb seals from greater dis- tances at Puffin Island. To test this, we used a theodolite to determine distance between seals and an approaching ves- sel at Puffin Island and Clements Reef There was, however, no significant (P>0.05) difference in distances at which disturbances occurred. The most notable difference in distance of distur- bance was between initial and subse- quent harassments during a haul-out period. Those seals remaining or re- turning to shore after a harassment were more tolerant of powerboats, al- lowing significantly (P<0.05 ) closer ap- proaches than those initially harassed. Seals detected (head raised and ori- ented toward the potential disturbance) a powerboat at a mean distance of 264 m, and harassments occurred when boats approached, on average, to within 144 m. Results of this study exemplify the variability in reaction to distur- bance and the necessity for consider- ing these differences for minimizing disturbance. Variability in reactions of Pacific harbor seals, Phoca vitulina richiardsi, to disturbance Robert M. Suryan James T. Harvey Moss Landing Marine Laboratories PO, Box 450 Moss Landing, California 95039 E-mail address (for R M Suryan) roberlsuryan li mail fws gov Present address (for R. M. Suryan): Migratory Bird Management U.S Fish and Wildlife Service 101 1 E Tudor Rd Anchorage, Alaska 99503 Manuscript accepted 26 may 1998. Fish. Bull. 97: 332-;«9 ( 1999). In many locations, disturbance is an important factor affecting the haul- out patterns of harbor seals, Phoca vitulina. Disturbance is defined as any activity that alters normal be- havior. In the United States, distur- bance of marine mammals by hu- mans is regulated by the Marine Mammal Protection Act of 1972. In contrast to pelagic marine mam- mals, changes in the behavior of pinnipeds on haul-out sites related to distLU'bance is relatively simple to measure. Long-term effects of disturbance, however, are often dif- ficult to assess. The effects of disturbance may be quite mild or may cause displace- ment and even mortality. Bighorn sheep (Ovis canadensis) and white- tailed deer exposed to snowmobile traffic have shown increased heart rate but no visible change in behav- ior (MacArthur et al., 1982; Moen et al., 1982). Humpback whale (Megaptera novaeangUae) female- calf pairs in Hawaii have avoided nearshore areas of intense human recreational activities (Salden, 1988; Glockner-Ferrari and FenariM, Dis- turbance-related mortality in har- bor seals can result from stamped- ing and pup abandonment (John- son'^). Disturbance from low-flying aircraft may have caused mortality of more than 200 ( IG'fi ) harbor seal pups on Tugidak Island, Alaska, in 1976 (Johnson-). In addition to aircraft, sources of disturbance include boats, seismic exploration, pedestrians, kayakers, and natural predators (Renouf et al, 1981; Laursen, 1982; Allen etal., 1984; Terhune, 1985; Richardson et al., 1995; Moss, 1992; Kroll, 1993; Johnson-; Murphy and Hoover^; Calambokidis et al."'; and others). Allen et al. (1984) reported that harbor seals on a haul-out site in Bolinas Lagoon, California were dis- turbed by humans on 71'7f of survey days; most disturbances were caused by nonmotorized boats ( primarily ca- noes ). Humans, primarily boat opera- tors, were the most common cause of harassment to harbor seals on Pro- ' Glockner-Ferrari, D. A., and M. J. Ferrari. 198.5. Individual identification, behavior, reproduction, and distribution of hump- back whales, Megaptera novacangliae, in Hawaii. Rep MMC-8.3/06 for Mar Mamni. Comm., 42 p. |NTIS PB85-200772.J - Johnson, B. W. 1977. The effects of hu- man disturbance on a population of har- bor seals. In Environmental assessment of the Alaskan continental shelf p. 422- 432. Annual Rep. Princ. Invest., vol. 1. U.S. Dep. Commer, NOAA/OCSEAP, 708 p. INTISPB-280934/l.l ' Murphy, E. C, and A. A. Hoover. 1981. Research study of the reactions of wildlife to boating activity along Kenai fjords coastline. Final Rep. to Nat. Park Serv., Anchorage, AK. 12,5 p. ^ Calambokidis, J., G. H. Steiger, J. R. Evans, and S. J. Jeffries. 1991. Cen- suses and disturbance of harbor seals at Woodard Bay and recommendations for protection. Final report to Washington Dep. Nat. Resources, Olympia. WA. 45 p. Suryan and Harvey: Variability in reaction to disturbance among Phoca vitulina richardsi 333 ^ r-i^' obs. site Skipjack Is. a^, ^ tection Island (Kroll, 1993) and Woodard Bay (Calambokidis et al."*), Washington. Sources of dis- turbance to harbor seals ashore at Gertioide Island, Washington, were mainly unidentifiable; how- ever, of detectable causes, human activities and coyotes were the most common (Moss, 1992). Reaction to disturbance may vary among harbor seal groups within an area (Terhune and Almon, 1983) and according to disturbance sources (e.g. power- boats vs. canoes and kayaks; Calambokidis et al.'*). This vari- ability may be attributed to dif- ferent levels of tolerance among age, sex, or reproductive status of harbor seals. Reaction to dif- ferent causes of disturbance may vary with exposure to particular sources, eventually resulting in greater avoidance or tolerance. In any case, results of previous stud- ies indicate that reaction to dis- turbances vary within and among regions, although little quantita- tive evidence exists. In this study, we collected data to evaluate the extent of disturbance to harbor seals at haul-out sites in the northern San Juan Islands. Our objectives were to determine 1 ) if human-related activities were the primary source of disturbance; 2 ) if recovery var- ied between flood and ebb tides and was similar among the three haul-out sites (one location was a pupping area); 3) if vigilance characteristics differed among haul-out sites; 4) if the response to harass- ment was similar for pups and for adults and sub- adults; and 5) if the mean distance between harbor seals and a boat causing a disturbance varied within and among haul-out sites and, if so, to determine potential causes of this variability. Methods Study area The study area was located in the northern San Juan Islands, Washington (Fig. 1 ). This area is character- ized by numerous islands, a tidal range of 3.6 m, strong currents (maximum of 7.7 km/h), and a rocky shore- line. Haul-out sites of harbor seals, which include reefs and rockv intertidal zones of islands, are numerous but 123° 00 W %/t o/-^ ^O/i 2Va 22° 50 W t 48° 50' N 0 N 5 Kilometers Patos Is. ^, „ ^ Clements Reef Sucia Is.nf^-^^^-s obs. sites \ "^ 5<^^^^ Ewing Is Matia Is 48° 45 N - Puffin Is. Figure 1 Locations (*) of harbor seal haul-out sites observed at Clements Reef, Puffin Island, and Skipjack Island during 1991 and 1992 pupping seasons in the San Juan Islands, Wash- ington. Dots ( • ) indicate other haul-out sites in the area. typically are used by fewer than 100 animals. During this study, observations were conducted at Clements Reef, Puffin Island, and Skipjack Island (Fig. 1). Observations of haul-out sites Ground-based surveys of harbor seals were conducted at Clements Reef (« = 13), Puffin Island («=9), and SkipjackIsland(n=8)from2 July to 19 August 1991. In 1992, surveys were conducted between 24 June and 10 September 1992 at Clements Reef (/2=21) and Puffin Island (n-l8). Skipjack Island was not surveyed during 1992 to allow increased sampling effort at the other two sites. Seals at each location were observed at least twice a week (one observer per site). Seals at Clements Reef were viewed ft-om Ewing Island (Fig. 1), approximately 0.55 km away. Seals on the north- west end of Puffin Island were viewed ft-om the south- east corner of Matia Island, 0.38 km away (Fig. 1). Observer heights above zero tide level were 10 m at Clements Reef and 13 m at Matia Island. The observa- tion point on the northeast side of Skipjack Island was directly above (23 m) the haul-out site (Fig. 1). Observations of harbor seals began one to three hours before low tide and ended three to seven hours after low tide (when <50% of the maximum number of seals counted during that tide cycle remained ashore). 334 Fishery Bulletin 97(2), 1999 Seals were viewed with 22x and 15-60x spotting scopes. Scan surveys ( Altmann, 1974) were conducted at ten minute intervals throughout the observation period. During each half-hour period, the first scan was a count of all seals, the second scan included size structure (number of harbor seal pups and sub- adults and adults), and the third scan was used to quantify vigilance of seals (head up, alert but not oriented toward a disturbance source) and sources of disturbance (within 1.0 km of the haul-out site). Counts of size structure included only those individu- als that could be assigned a given category (e.g. seals were not included if only a flipper was visible as in the first scan). Counts from these scans were used to determine how many seals entered the water following a dis- turbance and when recovery had occurred. Recovery was measured by the increase in number of harbor seals on the haul-out site after harassment. Recov- ery was divided into four categories: 1) full recovery (number of seals ashore after harassment returned to preharassment levels); 2) partial recovery (num- ber of seals ashore increased after the harassment, but did not reach preharassment levels); 3) no recov- ery (number of seals ashore did not increase after harassment); and 4) no chance to recover (number of seals ashore never increased after harassment owing to repeated disturbances or rising tide washing over the haul-out site). Because animals were not marked, full recovery did not imply that individuals returning to shore were necessarily the same ones that were ha- rassed, but partial and no recovery did indicate cer- tain individuals did not return to the haul-out site. Terhune (1985) and da Silva and Terhune (1988) reported that the number of vigilant harbor seals was dependent on group size. To eliminate the potential effect of group size on vigilance, the original data were subsampled to produce subsets of equal group sizes. Therefore, a single value is not presented for a site because it varies with each comparison. Data were collected for every potential source of disturbance that approached the haul-out site. Sources were divided into the following categories; airplanes, powerboats (including sailboats under motor power), sailboats, kayaks and canoes, people, bald eagles (Haliaeetus leucocephalus), unknown, or other Vessel speed was classified as underway fast (creating a breaking bow wake), underway slow (nonbreaking bow wake), and drifting (motor not in gear or turned off). Harbor seal reactions to a distur- bance were categorized as 1) detection: >1 seal with head raised and oriented toward potential distur- bance source; 2) alarmed: >1 seal moved from its rest- ing place, but did not enter the water; and 3) ha- rassed: >1 seal entered the water. Positions of an approaching vessel were monitored by using a Nikon NT2A or Pentax TH20D theodo- lite. Bearings to the approaching vessel and seals exhibiting disturbance reactions were recorded. The distance from theodolite to vessel or seal was calcu- lated by using the tangent of the vertical angle from the theodolite and height of the theodolite above the water. The distance between vessel and seals was calculated by using the Law of Cosines that incorpo- rates distances between theodolite and vessel and theodolite and seals and by using the horizontal angle between the vessel and seals (from theodolite). Height of the theodolite above water was measured directly or estimated from a cosine prediction of tide height ( San Juan Current and Tide Tables, published by Island Canoe, Bainbridge Island, Washington). The tidal constituent used was Port Townsend, Wash- ington, with a correction for Echo Bay, Sucia Island (approximately 1 km from Ewing Island and 6 km from Puffin Island). The observation point at Ewing Island (for Clements Reef surveys) was near a verti- cal rock ledge, which allowed the observer to mea- sure theodolite heights above water level (using a tape measure with float attached). Direct measure- ment of height above water was accurate to approxi- mately 0.1 m. Direct measurement was not possible at Puffin Island; therefore, theodolite heights above water level were based on tide height predictions, which were accurate to approximately ±0.3 m. Error in theodolite locations was less than 10 m; accuracy was based on calibration with fixed objects (e.g. buoy or island). Analyses Results of statistical analyses were considered sig- nificant at a = 0.05. Mean percentage of pups among the sites was compared by using analysis of variance ( ANOVA) with arcsine transformation. A Tukey mul- tiple comparison test was usedfor significant ANOVA results (Zar, 1984; Day and Quinn, 1989). To test whether pups were less tolerant of disturbance than adults, the frequency of positive and negative changes in the proportion of pups ashore before and after a harassment was compared by using a chi-square goodness-of-fit test (data were combined from all sites). Percentage of vigilant seals was compared between sites by using Mann- Whitney U tests (sepa- rate two-sample tests were conducted owing to ran- dom subsampling to control for group size). Differ- ences in distance of disturbance among powerboat approach speeds were tested with Mann-Whitney, t, and Kruskal-Wallis tests. Significant differences in distance of disturbance among categories of detec- tion, alarm, and harassment were detected with Suryan and Harvey: Variability in reaction to disturbance among Phoca vitulina nchardsi 335 Table 1 Summary of numbers of harbor seals using haul-out sites and primary dis turbances occurring during the pupping season of 1991 and 1992 in the northern San Juan Islands Washington. Puffin Island Clements Reef Skipjack Island No. of seals (range) 50-125 125-275 50-125 V, pups ( SE 1 17% (1.1) 3% (0.3) 3% (0.4) % of days with harassment (n ) 77% (27) 71% (34) 88% (8) No. of harassments observed 64 91 24 '7< of harassments caused by powerboats 42% 76% 46% Unknown' 19% 2% 38% Bald eagle 16% 2% 0% ' No disturbance source was detectable by the observer. ANOVA with square root transformation. Initial ver- sus subsequent disturbances were compared within each disturbance categories with ^tests. Randomiza- tion statistics ( Resamphng Stats, Inc., 1995 ) were used in calculation of power to detect significant differences. Results Clements Reef was the largest haul-out site with the greatest number of seals. Puffin Island, however, had the greatest number of pups, the pups representing a significantly (F=97.6, P<0.01) greater percentage of the total number of seals ashore in comparison with Clements Reef (Table 1 ).The study area on Puf- fin Island appeared to be an important area for fe- male-and-pup pairs and thus may explain some of the observed differences in reactions to disturbance. Harassments of seals occurred during at least 71'7f of the survey days. The primary cause of disturbance was powerboats (Table 1). Most of these (749^, n=96) involved boat operators approaching haul-out sites to view seals. The second most common category of disturbance was unknown (no source was visible or audible to the observer) and was most prevalent at Skipjack Island. Bald eagles were the third most com- mon cause of disturbance and were more frequent at Puffin Island than at other sites. In contrast with powerboats, sea kayaks were uncommon during the study and, therefore, caused fewer harassments ( 1 1% for all sites). Sea kayakers, however, were a greater potential disturbance to seals ashore than operators of powerboats. Fifty-five percent of kayakers {n=ll) within 1 km of a haul-out site harassed harbor seals, whereas only 9'7c of powerboats («=436) within 1 km caused harassment. This finding is a result of seals being less tolerant of kayaks and kayakers who gen- erally travel closer to shore than do powerboaters. Table 2 Percentage of harassment;: (/! . excluding disturbances caused by bald eagles I that resulted in either full recovery, partial | recovery, no recovery, or no chance to recover for harbor seals at Puffin Island, Clements Reef, and Skipjack Is and, north- em San Juan Islands, Wa shington, 1991 and 1992. Puffin Clements Skipjack Island Reef Island (n=27) (n=28) (n=ll) Full recovery 19% 54% 45% Partial recovery 30% 32% 18% No recovery 44% 7% 36% No chance to recover 7% 7% 0 For all recoveries no. of harassments before low tide 10 11 5 no. of harrassments after low tide 17 17 6 The extent of recovery following harassments (ex- cluding those caused by eagles) of harbor seals was less at Puffin Island than at Clements Reef and Skip- jack Island (Table 2). Disturbances caused by eagles were excluded from this comparison because eagles were potential predators of pups (eagles were ob- served approaching and disturbing pups) and could affect recovery. Recovery was related to whether the harassment occurred before or after low tide (74% of full recoveries occurred before low tide, whereas 75% of partial and 89% of no recoveries occurred after low tide). This finding corresponds with total num- bers of seals ashore typically decreasing one to two hours after low tide, independent of disturbance. For comparison of recovery among sites, the proportion of disturbances occurring before and after low tide was similar and likely did not influence results (Table 2). 336 Fishery Bulletin 97(2), 1999 350-1 Before Initial Harassment Alter Initial Harassment Deteelion Alarm Harassment Figure 2 Mean distance (+SEl between harbor seals and powerboat for disturbances occurring before and after the initial harassment during a haul-out period. An asterisk (*) indi- cates a significant (P<0.05) difference between before and after initial harassment data for that disturbance category. There was no apparent difference in the propor- tion of harbor seal pups onshore before and after harassments (x"=0.183, n=27, P>0.50). Pups did not appear to be affected disproportionately to subadults and adults. The percentage of vigilant harbor seals varied with each comparison (due to random subsampling of data to control for group size), hence there was no single value for percentage of vigilant harbor seals at each haul-out site. The percentage of vigilant harbor seals was significantly greater at Puffin Island compared with Clements Reef (P=0. 003) and Skipjack Island (P=0.002; Table 3). This finding is likely due to the greater percentage of female-and-pup pairs at Puf- fin Island and, in addition to lower recovery, indi- cated seals were more susceptible to disturbance. Powerboat speed did not significantly infiuence distance of disturbance. Lack of statistical signifi- cance was likely due to small sample sizes after sub- dividing data by approach speed and initial versus subsequent harassments. There was also no signifi- cant difference in distance of .disturbance between sites (data were not collected at Skipjack Island). Given the large variability of the data and small ef- fect size, greater than 450 observations would be re- quired for a power of 0.80 to detect a significant dif- ference between sites. Data, therefore, were pooled for further analysis. Twenty-five percent of harass- ments occurred when vessels were < 100 m from seals, 50% occurred at 100-200 m, and 25% at 200-300 m. After detection by seals, powerboats were able to approach significantly closer (P=10. 51, P<0.001) be- fore causing alarm or harassment of seals (Fig. 2). Interestingly, we found that distances of initial dis- turbance were significantly greater than subsequent disturbances (Fig. 2). This indicated that seals re- maining or returning to shore following the first ha- rassment were less easily disturbed. Discussion Many islands in the San Juan Archipelago are state parks or have resort harbors that attract numerous boaters during the summer. Clements Reef is located near (0.6 km) Sucia Island, which is the most heavily visited state park in the northern islands. It was not surprising, therefore, that most disturbances at Clements Reef were caused by boaters. Relatively few kayakers ventured out to Clements Reef, Puffin Island, or Skipjack Island. Kayakers typically travel along the shoreline and have been shown to cause Suryan and Harvey: Variability in reaction to disturbance among Phoca vitulina nchardsl 337 Table 3 Mean percentage of vigilant harbor seals at Puffin Is and versus Clements Reef and at Puffin Island versus Skipjack Is land when no potent ial distL rbance source was within 1 km of haul-out site. Two means ( ±SE) and separate statistical tests are presented for Puffin Island due to random subsampling to control for group size of each comparison. Mean (%) SE n Mann-Whitney U P Puffin Island 4.43 0.44 17 Clements Reef 2.41 0.47 17 U=229 0.003 Puffin Isl and 6.40 1.13 21 Slipjack I sland 2.46 0.75 21 U=343 0.002 harassment of harbor seals at a greater distance than do operators of powerboats (Calambokidis et al.''); therefore, as sea kayaking becomes more popular, there is a greater potential for disturbance of harbor seals ashore. The differences in occurrence of unknown causes of disturbance among haul-out sites was possibly a result of haul-out site topography. Harbor seals on Clements Reef had a 360° view of potential sources of disturbance compared with roughly a 270° view for Puffm Island and 180° for Skipjack Island. A dis- turbance of unknown origin at Skipjack Island would often begin by several harbor seals looking toward the rocky cliff of the island, then entering the water. A high incidence of disturbances of unknown origin have also been documented at Gertrude Island ( 7T7( ; Moss, 1992) and Protection Island (43%; Kroll, 1993), Washington. The relatively high occurrence of ha- rassments by bald eagles at Puffin Island may have been due to a nearby bald eagle nest and the high percentage of harbor seal pups (immature eagles were observed harassing female and pup pairs ). Skip- jack Island also had an active bald eagle nest, but eagles were not observed harassing harbor seals. Overall, only 39% of all harassments resulted in full recovery, indicating seals often remained in the water or moved to a different site. Allen et al. ( 1984) reported that the number of harbor seals that re- turned to a haul-out site after a disturbance in Bolinas Lagoon, California, was always less than the original number, and in most cases, harbor seals did not move to a nearby reef Murphy and Hoover"^ re- ported that harbor seals off the Kenai Qords, Alaska, often searched for a new haul-out site after harass- ment. Disturbance to harbor seals, therefore, may have considerable impact where haul-out space is limited (Murphy and Hoover'^). Although haul-out sites in the San Juan Islands are numerous, alter- nate sites for female and pup pairs, similar to Puffin Island, may not be readily accessible (particularly since there tended to be less recovery of seals at Puf- fin Island). Terhune (1985) compared aggregation behavior and vigilance of harbor seals with flocking behavior of avian species — a behavior that allows individuals to decrease their surveillance without decreasing the probability of detecting a predator (Caraco, 1979; Studd et al., 1983). Da Silva and Terhune ( 1988) iden- tified group size as the only factor accounting for variation in time taken to scan for predators. Renouf and Lawson (1986) suggested that only males in- creased scanning time as mating season approached, and scans were related to important events in their mating system, not predators. Results of our study indicated that increased vigilance may be related more to potential "predators" (loosely defined as any source of disturbance, human or animal). We found that seals at an area with a greater percentage of female-and-pup pairs scanned more frequently than those at other locations. Other researchers have de- scribed increased vigilance of females with pups. Stein ( 1989 ) reported female harbor seals rested alert significantly more frequently when their pups were one to nine days old than when pups were older Newby (1973) reported a female harbor seal with a pup is "constantly alert and nervous." The greater vigilance of harbor seals at Puffin Is- land than at Clements Reef and Skipjack Island and the lack of recovery from a harassment indicated that seals at a pupping location were affected more by disturbance. We therefore expected that seals at Puffin Island would enter the water when powerboats were farther away, in contrast with harbor seals at Clements Reef. This was not the case, there were no significant diff"erences between sites. We cannot, how- ever, conclude that seals at pupping locations toler- ated boats to approach just as closely as seals at nonpupping sites without harassment because of dif- ferences in geographic characteristics of haul-out sites, unreplicated sites, and lack of power to detect a statistically significant difference. Bishop (1967) observed that a nucleus of harbor seals, which usually included several very large ani- mals, remained ashore unless danger became immi- 338 Fishery Bulletin 97(2), 1999 nent. Although we also observed certain groups of harbor seals (often small-size groups) entering the water well before others during a harassment, there was no significant trend. Terhune and Almon (1983) also noted that not all groups of harbor seals reacted to disturbances in the same manner. The most notable difference in distance of distur- bance was the decrease between initial and subse- quent harassments. There are two plausible expla- nations for this: 1) seals became more tolerant of powerboat approaches; or 2) seals that were less tol- erant of disturbance did not return to the haul-out site after the initial harassment. Based on lack of full recovery following initial harassments, our re- sults support the latter possibility. Allen et al. (1984) reported that boats advancing toward, or remaining near, harbor seals ashore at Bolinas Lagoon, California, caused seals to leave the haul-out site more often than a boat simply moving past seals. Boats that traveled slowly, parallel to the haul-out site, and made no abrupt move or changes in speed approached harbor seals at Clements Reef and Puffin Island with minimal disturbance. Greater sample sizes probably would have resulted in sig- nificant differences among distances of disturbance for various approach speeds and angles recorded during this study. Additional potential factors affecting the distance at which seals were disturbed included time of day for haul-out period or location of haul-out site. Har- bor seals may more readily enter the water toward the end of the haul-out period, if air temperature is high (Watts, 1992), or during rain. These consider- ations were not addressed in this study Harbor seals also may become accustomed to close approaches ( < 15 m) by boats in areas of high boat traffic (authors' pers. obs.). In Woodard Bay, Calambokidis et al."* re- ported powerboats were able to approach to an aver- age of 40 m before harbor seals entered the water. This area near Seattle, Washington, undoubtedly gets more year-around recreational traffic than the northern San Juan Islands which may explain why disturbance distances were over a third less than those observed during our study Most harassments in our study were caused by people approaching to view harbor seals. It is impor- tant therefore to distribute information and guide- lines for wildlife viewing to the general public. An effective solution may be to include information with vessel registration. Direct mortality of harbor seals or long-term aban- donment of haul-out sites (greater than one haul- out period) because of harassment by humans was not obsen-'ed during this study. Long-term impacts of harassment of harbor seal populations, however. are difficult to assess. Cases where marine mammals remain in heavily disturbed areas are easy to detect; therefore, cases of partial or complete abandonment of disturbed areas may be more common than evi- dence indicates (Richardson et al., 1995). Harass- ments increase energy expenditure of harbor seals by decreasing duration of haul-out period. Increased energy requirements likely have the greatest impact on harbor seal pups during nursing and on adult and subadult harbor seals during molt when access to haul-out sites is important. Brasseur et al. (1996), however, demonstrated that captive harbor seals may need to haul-out even outside these "critical" peri- ods. Richardson et al. (1995) noted that occasional disturbance probably has little effect on harbor seal populations; repeated disturbance, however, may have significant negative effects especially at haul- out sites used for pup rearing. Results of this study quantify the variability in response to disturbance among individuals and lo- cations. We also demonstrated the potential bias in sampling animals that remain after an initial dis- turbance, or in areas of regular boat traffic. Distance at which powerboats caused harassment were vari- able, ranging from 28 m to 260 m. Boating regula- tions near harbor seal haul-out sites should address activity of vessel (speed and approach angle) in ad- dition to distance from harbor seals. Boating precau- tions are particularly important near harbor seal pupping areas such as Puffin Island. Acknowledgments We thank Birgit Kriete with The Whale Museum and Robert DeLong with the National Marine Mammal Laboratory for providing initial assistance and ac- quiring funding to conduct field work in 1991. Steve Jeffries and Harriet Huber provided equipment and assistance during both field seasons. Kim Raum- Suryan was instrumental in data collection and lo- gistics for conducting field work. For assistance in the field we thank Karen Russell, Tomo Eguchi, and Doug Huddle. Additional assistance was provided by Steve Osmek and David Rugh. Robert DeLong, Michael Foster, and two anonymous reviewers pro- vided critical review of early versions of this manu- script. John Calambokidis and Peter Watts provided helpful discussions on studying disturbance of har- bor seals. Primary funding was provided by the Wash- ington Department of Wildlife and the National Marine Mammal Laboratory. Additional funding was provided by the Earl H. and Ethel M. Myers' Oceano- graphic and Marine Biology Trust "Save the Whales, Inc.," and the Packard Foundation. This study was Suryan and Harvey: Variability in reaction to disturbance among Phoca vttuhna nchardst 339 conducted under Marine Mammal Protection Act permit #473 issued to the Washington Department of Wildhfe for scientific research. Literature cited Allen, S. G., D. G. Ainley, and G. W. Page. 1984. The effect of disturbance on harbor seal haul out pat- terns at Bolinas Lagoon, California. Fish. Bull. 82( 3 ):493- 500. Altmann, J. 1974. Observational study of behavior: sampling methods. Behavior 49:227-267. Bishop. R. H. 1967, Reproduction, age determination, and behavior of the harbor seal, Phoca citulina I. in the Gulf of Alaska. M.S. thesis, Univ. Alaska, Fairbanks. AK, 121 p. Brasseur, S., J. Creuwels, B. v/d Werf, and P. Reijnders. 1996. Deprivation indicates necessity for haul-out in har- bor seals. Mar. Mamm. Sci. 12(4):619-624. Caraco, T. 1979. Time budgetmg and group size: a test of the theory. Ecol. 60:618-627. da Silva, J., and J. M. Terhune. 1988. Harbor seal grouping as an anti-predator strategy. Anim. Behav 36:1309-1316. Day, R. W., and G. P. Quinn. 1989. Comparisons of treatments after an analysis of vari- ance in ecology. Ecol. Monogr. 59(41:433-463. Kroll, A. M. 1993. Haul out patterns and behaviorof harbor seals, Phoca vitulina. during the breeding season at Protection Island, Washington. M.S. thesis, Univ. Washington, Seattle, WA, 142 p. Laursen, K. 1982. Recreational activities and wildlife aspects in the Danish Wadden Sea. Schriftenreihe des undesministers fiir Ernahrung. Landwirtschaft und Forsten 275:63-83. MacArthur, R. A., V. Geist, and R. H. Johnston. 1982. Cardiac and behavioral responses of mountain sheep to human disturbance. .J. Wildl, Manage. 46(21:351-358. Moen, A. N., S. Whittemore, and B. Buxton. 1982. Effects of disturbance by snowmobiles on heart rate of captive white-tailed deer. N.Y. Fish and Game J. 29(2):176-183. Moss, J. 1992. Environmental and biological factors that influence harbor seal iPhoca citulina richardsi) haul out behavior in Washington and their consequences for the design of population surveys. M.S. thesis. Univ. Washington, Se- attle, WA, 122 p. Newby, T. C. 1973. Observations on the breeding behavior of the harbor seal in the State of Washington. J. Mammal. 54:540-543. Renouf, D. and J. W. Lawson. 1986. Harbor seal I'igilance: watching for predators or mates? Biol. Behav 11:44-49. Renouf, D., L. Gaborka, G. Galway, and R. Finlayson. 1981. The effect of disturbance on the daily movements of harbour seals and grey seals between the sea and their haul- ing grounds at Miquelon. Appl. Anim. Ethol. 7:373-379. Resampling Stats, Inc. 1995. Resampling stats users guide. Resampling Stats, Inc., Arlington, VA. 128 p. Richardson, W. J., C. R. Greene Jr., C. I. Malme, and D. H. Thomson. 1995. Marine mammals and noise. Academic Press, Inc. San Diego. CA, 576 p. Salden, D. R. 1988. Humpback whale encounter rates offshore of Maui, Hawaii. J. WUdl. Manage. 52(21:301-304. Stein, J. L. 1989. Reproductive parameters and behavior of mother and pup harbor seals, Phoca vitulina richardsi, in Grays Har- bor, Washington. M.S. thesis, San Francisco State Univ., San Francisco, CA, 110 p. Studd, M., R. D. Montgomerie, and R. J. Robertson. 1983. Group size and predator surveillance in foraging house sparrows {Passer domesticus). Can. J. Zool. 61: 226-231. Terhune, J. M. 1985. Scanning behavior of harbor seals on haul-out sites. J. Mamm. 62(2):392-395. Terhune, J. M. and M. Almon. 1983. Variability of harbour seal numbers on haul-out sites. Aquat. Mamm. 10:71-78. Watts, P 1992. Thermal constraints on hauling out by harbour seals iPhoca citulma ). Can. J. Zool. 70:553-560. Zar, J. H. 1984. Biostatistical analysis. Prentice-Hall, Englewood Cliffs, NJ, 718 p. 340 Abstract.— Age structure, longevity, and mortality were determined for a population of blackcheek tonguefish, Symphurus plagiusa. in Chesapeake Bay. Blackcheek tonguefish (36-202 mm TL) were randomly collected by means of otter trawl in lower Chesa- peake Bay and major Virginia tributar- ies (James. York, and Rappahannock rivers) from April 1994 through August 1995. Ages were determined by inter- preting growth increments on thin transverse sections of sagittal otoliths. Marginal increment analysis showed that a single annulus was formed in June of each year. Blackcheek tongue- fish caught during this study reached a maximum age of 5-i- years. Growth was rapid during the first year, then slowed rapidly at a time coincident with maturation. We used the following von Bertalanffv growth equations: for males— L, = 196.5(1 -e-«^«5(, + 0.92.). anj for females— L, = 190.6(1 -e-^^^"" *'"*"). Von Bertalanffy parameters were not significantly different between sexes. Extrapolated instantaneous mortality rates for a possible seventh year class were 0.73 ( Hoenig's equation i and 0.77 (Royce's equation I. High estimates of instantaneous total mortality may re- flect either loss due to emigration of adults from Chesapeake Bay onto the continental shelf or high natural mor- tality occurring in this northernmost population. Compared with sympatric pleuronectiforms, blackcheek tongue- fish have a relatively high mortality rate, small asymptotic length, and high growth parameter A'. Species, such as blackcheek tonguefish. that feature this combination of growth parameters are hypothesized to be better adapted at exploiting seasonally dynamic and highly unpredictable environments, such as those estuarine habitats within Chesapeake Bay. Age, growth, longevity, and mortality of blackcheek tonguefish, Symphurus plagiusa (Cynoglossidae: Pleuronectiformes), in Chesapeake Bay, Virginia* Mark R. Terwilliger School of Marine Science, Virginia Institute of Marine Science College of William and Mary Gloucester Point, Virginia 23062 Present address: 104 Nash Hall Department of Fisheries and Wildlife Oregon State University Corvallis, Oregon 97331 E-mail address terwillm d ucs orst edu Thomas A. Munroe National Marine Fisheries Service National Systematics Laboratory, MRC 153 National Museum of Natural History Washington, DC. 20560 Manuscript accepted 15 June 1998, Fish. Bull. 97:340-.361 (1999). The blackcheek tonguefish, Sym- phurus plagiusa (Linnaeus, 1766), ranges from Connecticut southward through the Florida Keys, northern Gulf of Mexico, Cuba, and the Ba- hamas (Ginsburg, 1951; Topp and Hoff, 1972; Munroe, 1998). Chesa- peake Bay is the northernmost lo- cation of a major population of this species (Munroe, 1998). South of Chesapeake Bay, blackcheek tongue- fishes are an abundant component of the fish fauna occurring in estuaries and inshore coastal waters. Within this region, they occur in sufficient numbers to form a minor component of the bycatch of demersal fisheries and also constitute a minor inclu- sion in landings reported for several industrial fisheries. In the shrimp trawl fishery, they are a potential nuisance because individuals fre- quently become embedded in the webbing of trawls to such an extent that they interfere with the gear's effectiveness (Topp and Hoff 1972). The blackcheek tonguefish is the only member of the pleuronectiform family Cynoglossidae occurring in Chesapeake Bay (Ginsburg, 1951; Murdy et al., 1997; Munroe, 1998), This species is among the top three most abundant pleuronectiforms occurring throughout lower Chesa- peake Bay and its tributaries (Bonzek et al., 1993; Geer et al,, 1997 ). Only the hogchoker, Trinectes maculatus, is commonly taken in greater abundance in bottom trawls made within the Bay. Occurrence of all life history stages in Chesapeake Bay (Olney and Grant, 1976; Ter- williger, 1996) suggests that black- cheek tonguefish is a resident spe- cies in this system. Despite its abundance in differ- ent estuarine and nearshore marine habitats in the northwest Atlantic, life history parameters of S. pla- giusa are largely unknown. Some previous works have described dis- tribution, relative abundances and length frequencies for blackcheek tonguefish in South Carolina estu- aries (Shealy et al., 1974); some have summarized size-related ma- turity patterns for individuals from throughout the entire range of the species (Munroe, 1998); some have ' Contribution 2203 of the Virginia Institute of Marine Science, College of William and Marv Gloucester Point. VA 23062. Terwilliger and Munroe: Age, growth, longevity, and mortality of Symphurus plagiusa 341 investigated factors influencing recruitment (Miller et al., 1991); and some have described daily growth rates of juveniles in Georgia estuaries (Reichert and van der Veer, 1991). The minimal amount of pub- lished information on this species (summarized in Munroe, 1998) may be due in part to the relatively small size of the fish. It reaches a maximum size of approximately 210 mm total length (TL) (Wenner and Sedberry, 1989), but fish smaller than 165 mm TL are those most commonly taken in Chesapeake Bay by otter trawl (Terwilliger, 1996; Geer et al., 1997). Small size and benthic microinvertebrate feeding habits (Stickney, 1976; Reichert and van der Veer, 1991; Toepfer and Fleeger, 1995) render this species inaccessible to most recreational and commercial [l fishing gears used in Chesapeake Bay. Few age and growth studies using bony structures or otoliths exist for species of the Cynoglossidae. This relatively large gap in knowledge for approximately 150 species of tonguefishes was recently noted in a compilation of flatfish life history parameters (Miller etal., 1991). Although several studies have described growth for species of Cynoglossiis from the eastern Atlantic (Chauvet, 1972), western Pacific (Lin, 1982; Meng and Ren, 1988; Zhu and Ma, 1992), and In- dian Ocean (Seshappa, 1976, 1978, 1981; Ramana- than et al., 1980; Seshappa and Chakrapani, 1984), no growth studies using bony structures or otoliths have been performed on species oi^ Symphurus. Pre- vious work describing the age structure of blackcheek tonguefish in Chesapeake Bay using length frequen- cies offish taken by otter trawl (Bonzek et al., 1993) is limited because this type of analysis requires sub- jective interpretation of modal frequencies in the data, which, given the difficulty of distinguishing modal groups at older ages, renders the interpreta- tion unreliable (Chauvet, 1972; White and Chitten- den, 1977; Jearld, 1983; Barbieri, 1993). Such limi- tations in interpretation of age from length-frequency distributions are particularly evident in data from Koski's ( 1978) study on hogchokers from the Hudson River estuary, where the first of bimodal peaks in a frequency distribution of hogchoker lengths corre- sponded to age-1 fish, whereas the second peak rep- resented combinations of length distributions for fishes age 2-6 yr. Chauvet's (1972) data for C. canarlensls off West Africa also show a similar pattern of overlapping sizes among fishes of different ages. Syr7iphurus plagiusa is a good candidate for an age and growth study. It is unique among the Cynoglos- sidae in that it is the only species of this family that inhabits estuarine environments in the seasonally dynamic region of north temperate latitudes. Most other cynoglossid species occur in the tropics and are difficult to age from hard parts because environmen- tal constancy precludes the formation of interpret- able annular growth marks on these structures (Qasim, 1973). Other species of symphurine tonguefishes that occur in temperate waters gener- ally are small-size, deep-water forms that are diffi- cult to catch in abundance (Munroe, 1998). This study was undertaken to determine age struc- ture, growth rate, longevity, and mortality for a popu- lation of S. plagiusa residing in Chesapeake Bay, Virginia. Knowledge about life history parameters for this species provides a window of understanding into the biology of the Cynoglossidae, as well as in- sights into age, growth, and longevity of other small- size, estuarine-dependent flatfishes. Materials and methods Blackcheek tonguefishes were collected by 9.14-m semiballoon otter trawl with a 38.10-mm stretch mesh body, 6.35-mm mesh codend liner, and attached tickler chain (Bonzek et al., 1993). Fish were collected from April 1994 through August 1995 during the Vir- ginia Institute of Marine Science (VIMS) Juvenile Finfish and Blue Crab Stock Assessment Program's trawl survey. The survey employs a monthly, ran- dom-stratified design of the lower Chesapeake Bay and fixed-station mid-channel transects in each of three major Virginia tributaries; the York, James, and Rappahannock rivers (Fig. 1). Details of sam- pling design were provided in Geer et al. (1993). Samples of blackcheek tonguefishes selected fi-om trawl catches were brought to the laboratory, measured for total length (TL) to the nearest millimeter and for total weight (TW) taken to the nearest hundredth of a gram. When samples were small, all fish were used in age analyses; however, fish were randomly selected for age analyses from relatively large samples. Regressions were fitted to the length-weight data, and regressions on log-transformed data for male and female blackcheek tonguefish were compared by using analysis of covari- ance (ANCOVA). The formula to convert SL to TL for blackcheek tonguefish provided by Jorgenson and Miller ( 1968: 1 1 ) was used when comparing length data (reported as SL) from other studies. All fish were sexed macroscopically, and then gonads were removed and preserved for later histological analy- sis to determine maturity stages (Terwilliger, 1996). Males were classified as either immature or mature according to the absence or presence of spermatozoa in the testes; females were considered mature if they showed any evidence of developing oocytes or previous spawning, e.g. thickening of the ovarian membrane. Both sagittal otoliths were removed, cleaned, and stored dry for later processing. Otolith maximum 342 Fishery Bulletin 97(2), 1999 75°30' 75 75^^0' .j^ M rJ^ -< ^^"* 38" .~'i £,- \ "'' "' y¥' / " \ K ■ ■""^'"■,. . V ■ '>*' ••'■^'l-;?- |. ^ f'-^' \ ■• N A, /'Y 1 J A •" ([ - ■■i "P- ^ T, • • & . • • • - # / '\/ '\ . % • .... -v. • 37° - . , ...... . 0 10 20 30 40 Kilometers _ 38° 37» 76° 75°30' Figure 1 Lower Chesapeake Bay system, including James, York, and Rappahannock rivers. Points indicate stations where blackcheek tonguefish, Symphurus plagiusa, were collected, April 1994— August 1995. diameter (distance from anterior tip of rostrum to postrostrum) was measured to the nearest 0.00001 mm by using a compound microscope coupled with a video camera interfaced with a microcomputer equipped with Biosonics' Optical Pattern Recognition System (OPRS) software (Biosonics, Inc., 1987). Be- cause no asymmetry in otolith shape was apparent, sagittal otoliths (left or right) were arbitrarily se- lected from each fish for age analysis, embedded in epoxy resin (Spurr, 1969), and sectioned transversely ( 1 mm thick ) through the core with a Buehler Isomet low-speed saw with duel diamond blades. Sections were mounted on glass slides with Crystal Bond ad- hesive, sanded with 1000-grit sandpaper to remove saw marks and to gain proximity to the core, pol- ished with alumina powder, and examined with a binocular dissecting microscope ( 30x ) with transmit- ted light and bright field. Annuli on otoliths from all 566 blackcheek tonguefish were counted once a month for three months. Counts were made without information regarding fish length or catch date. In 185 (32.7%) cases, the first and second readings did not agree and a third reading was made. The major- ity of disagreements in age estimates occurred dur- ing the beginning of the study; subsequent re-age- ing resolved most discrepancies in age estimates. In most cases, the second reading of an otolith section differed from the first reading by only one year If a third reading agreed with either of the first two, then that age was assigned to the otolith. In 28 (59f ) cases, the third reading was different than the first and second readings. These 28 otoliths were considered unreadable and excluded from further analysis. Otolith annuli were validated by the marginal in- crement method (Bagenal and Tesch, 1978; Jearld, 1983). Distances from the core to each annulus and the proximal edge were computed by drawing a ver- tical line from the core to the proximal edge of the Terwilliger and Munroe: Age, growth, longevity, and mortality of Symphurus plagiusa 343 otolith with the OPRS. The average margin width, i.e. the translucent region between the last annulus and the proximal edge of the otolith, was plotted by month. Regression analyses of otolith maximum di- ameter on total length and of weight on total length were calculated by the method of least squares. Back-calculated lengths-at-age were computed by using the Lee method (Lagler, 1956). To evaluate growth, back-calculated lengths-at-ages were fitted to the von Berta- lanffy model (Ricker, 1975) by using nonlin- ear regression ( Marquardt method ) calculated with Fishparm computer software (Saila et al., 1988). Likelihood-ratio tests were used to com- pare parameter estimates of the von Berta- lanffy equation for males and females (Ki- mura, 1980; Cerrato, 1990). A scale sample from a region in the poste- rior third of the body, dorsal to the midline, was taken from all fish. Six scales from each fish were cleaned with hydrogen peroxide and mounted on plastic strips by using a Carver hydraulic laboratory press. A random sample of scale strips from 50 fish were then viewed with a microfiche reader in order to discern annuli for comparison with otolith sections. In contrast to otoliths, scales proved to be unre- liable for ageing this species. Annuli were poorly defined and difficult to distinguish, and in many cases, the scales were unreadable. Fish ages obtained from scales agreed with ages obtained from otoliths in only 12% of the cases. This is not surprising, because scale annuli from other fish species with known ages have been shown to be inconsistent indicators of age, even in relatively young fish (Prather, 1967; Heidinger and Clodfelter, 1987). Instantaneous total annual mortality rates, Z, were estimated from maximum age esti- mates with a pooled regression equation as suggested by Hoenig (1983) and by calculat- ing a theoretical total mortality for the entire lifespan following the reasoning of Royce (1972) as described in Chittenden and McEachran ( 1976 ). Values of Z were converted to total annual mortality rates. A, using the relationship A = 1 -e"^ (Ricker, 1975). Results A total of 566 fish, encompassing a size range from 36 to 202 mm TL (Fig. 2A), were used in this ageing study. This size range included individuals from all c m 3 0.25). The length-weight relationship for sexes combined was W= 10 -.■>., 'i-i TL 3.1-)8 ). Size at sexual maturity observed in female and male blackcheek tonguefish ranged between 80-130 mm TL and 70-110 mm TL, respectively. Length at which 50% of the population reached maturity was 101 mm TL for females and 91 mm TL for males (Fig. 4). Sagittal otoliths of blackcheek tonguefish are small, round, dense structures that cannot be read whole. Transverse sections of otoliths (Fig. 5), how- ever, revealed fairly distinct opaque bands that could be counted. Transverse sections through otoliths in- dicated that otolith structure consists of an opaque core usually surrounded by a wide opaque area whose outer edge demarcates the first annulus. Size and appearance of this first annulus vary among fish, ranging from a broad band continuous with the core to a narrow opaque band not continuous with the core. Subsequent annuli are represented by thin opaque bands encircling the core and by broader translucent zones found between annuli. At the mar- gin of the otolith, an annulus was difficult to observe. However, when sufficient spring and early summer growth had occurred and contributed translucent material to the margin of the otolith after annulus Terwilliger and Munroe; Age, growth, longevity, and mortality of Symphurus plagiusa 345 90 J 80 -- 70-- 60-- S 50-- I 40-- 30 ■- 20 - 10-- 0 FEMALES: (f = 10"'" (7l"") MALES: W = XO'^'" (TL"^'^ ^^ ■ Males A Females 200 250 Total length (mm) Figure 3 Total length-weight relationship for both sexes of blackcheek tonguefish, Symphurus plagiusa, from lower Chesapeake Bay and major lower tributaries. formation, the densely compacted opaque zone delimiting the annulus was easily identifiable. The combination of one broad translucent zone and one narrow opaque zone (Fig. 5) represents one year's growth. Otolith maximum diameter (OMD) was linearly related to total length (Fig. 6) and described by the following relationship: TL = 64.59(OMD)- 17.02. [r2=0.93] This linear relationship indicated that otolith growth was proportional to fish growth. Size at first annulus formation for blackcheek tonguefish ranged from 88 to 138 mm TL (mean length ca. 118 mmTL). Seasonal progression of modal length frequencies (Fig. 7) for young-of-the-year fish (YOY) corroborated first-year growth-rate estimates determined from annular marks on the otoliths. Pooled data for 1993-96 from VIMS^ trawl survey reports indicated that YOY blackcheek tonguefish recruited to the gear during September, usually at sizes of ca. 35-45 mm TL. By the following June, these fish had reached sizes ranging from ca. 53 to 138 mm TL. Thus, empirical data from length fre- ' VIMS (Virginia Institute of Marine Science I. Department of Fisheries, Gloucester Point, VA 23062. Unpubl. data. quencies agreed fairly well with first-year growth estimates derived from interpretation of annulus for- mation on the otoliths. Monthly mean marginal increments on the otoliths were plotted for fish from all age groups (Fig. 8). The seasonal progression of marginal increments was similar on otoliths from all age groups; therefore, data were pooled to demonstrate general trends in annual growth. Marginal increments were smallest during June of both 1994 and 1995. Monthly mean marginal increments showed only one trough during a year, indicating that a single annulus is formed yearly in June. In 1994, margin width on the otoliths increased rapidly during July and August, reflecting an active period of growth. After August 1994, margin width leveled off and remained fairly constant from Sep- tember 1994 through May 1995. Large variation (in- dicated by high standard error) for samples taken during May of both 1994 and 1995 may indicate that the season of annulus formation may be somewhat more protracted (May-June) and perhaps dependent upon environmental factors. Alternatively, it may simply reflect an artifact of the small sample sizes examined from this period. Mean lengths at age were backcalculated for 92 males and 145 females (Table 2). Remaining fish were young-of-the-year and therefore were excluded from this analysis. Observed (empirical) lengths were con- 346 Fishery Bulletin 97(2), 1999 Iwl sistently higher than the back-calculated lengths-at- age for individual age groups, which indicated that seasonal growth had occurred since formation of a new annulus. Differences between back-calculated lengths-at-age and observed lengths are in the range of observed seasonal growth. Males and females have similar lengths-at-age until age 4 (Table 2). Beyond age 4, mean lengths for females were slightly greater than those for males. The largest male collected in this study was 190 mm (age 5+); the largest female collected was 202 mm (age 5h-). Greatest incremental growth in TL for both sexes occurred during the first year and then declined rap- idly thereafter (Table 2). Mean back-calculated lengths for females and males at the end of their first year were 78.84 mm and 75.68 mm, respectively. Growth for both sexes in the second year was only 1U0-] 90- ; A / 80- / • 70- / 60- « /• 40- « / 30- 1 20- ■ / 10' ■ y n^ c^ — 1- -^ L^- — (- — 1 — 1 — 1 — 1 — 1 — 1 — 1 — 1 — 1 ■•- ^' ^^^.,plH^ 50 40 30 20 10 -+— I — I — y A March m f -t — I— I — 1 — I o f 40j 3 30 ■• 20 ■• 10 ■■ 0 ■- April ilU 60 T 40 20 ■■ 0 llUMllllllf 60 40 20 0 -I — I — I — I — I — I — I- _ June Length (mm) Figure 7 Seasonal progression of modal length frequencies for blackcheek tonguefish, Symphurus plagiusa, from lower Chesapeake Bay and major lower tributaries, 1993-96. low), it also seems equally possible that the popula- tion of blackcheek tonguefishes residing in Chesa- peake Bay could be of more recent origin because all July H 1 1 1 1 I T — I — *- August -4 1 1 h— ( 1 1 h- JUIIU 150 100 ■■ 50 ■■ = 0 September nww -•r-m- E 200 ^ 150 ■■ 100 ■• 50 ■• 0 300 200 100 0 October -H 1 1 h- November -I — I — I — y llllln .llll. 200 150 100 50 •- H — I — I — h-^ uu December PTfTT¥¥¥-t-i I — I Length (mm) Figure 7 (continued) other major populations for this species are known to occur only in areas further to the south. If indeed this is the case, then life history parameters esti- mated for blackcheek tonguefishes from Chesapeake Bay are those characterizing a peripheral popula- tion of the species. Comparisons of these estimates 350 Fishery Bulletin 97(2), 1999 of life history parameters, then, with values derived from other, more centrally located, populations of this species, as well as with those from populations in- habiting subtropical environments in northern Mexico and Cuba (Munroe, 1998), would prove in- teresting from the perspective of discovering the range of variability in life history features that has evolved within this species of flatfish. Sizes of blackcheek tonguefishes collected within Chesapeake Bay during the present study represent almost the complete size range known for the spe- cies (Munroe, 1998). The largest female (202 mm TL) and largest male ( 190 mm TL) taken within the Bay miOmiotOlOlDlT) Q. < o z A 05 O) CT> a> O) o a> O) () (- j3 il il >, c 3 m 0 n! HI 5 ro 3 Q -~i LL < > —i < Date Figure 8 Plot of monthly mean marginal increments for blackcheek tonguefish, Symphurus plagiusa. ages 1-5, collected in lower Chesapeake Bay and ma- jor lower tributaries. Vertical bars represent ±1 standard error Numbers above bars represent monthly sample sizes. approach the largest sizes known for this species (ca. 210 mm TL; Wenner and Sedberry, 1989). However, most blackcheek tonguefishes occurring within this system were smaller, with total lengths usually rang- ing between 35 and 150 mm. The size range for blackcheek tonguefishes occurring in Chesapeake Bay is not unlike those reported for other estuarine populations located throughout the species' range. For example, Shealy et al. ( 1974) reported that, dur- ing a year-long survey offish assemblages in South Carolina estuaries, blackcheek tonguefishes ranged in size from 53 to 156 mm TL. However, modal lengths for this species were never greater than 140 mm TL, and fish were usually much smaller (monthly mean size 97—124 mm TL). In fact, of the blackcheek tonguefishes taken during that study, only 8 of 362 (2.2%) were 150 mm TL or larger. Although many studies (summa- rized in Munroe, 1998) have reported catching large blackcheek tonguefish from neritic habitats on the inner con- tinental shelf throughout the species' range, size ranges for these fishes were not unlike those observed for black- cheek tonguefishes collected within Chesapeake Bay. For example, black- cheek tonguefishes (n=439) collected at the mouth of Chesapeake Bay and on the inner continental shelf off south- ern Virginia and northern North Caro- lina (VIMS'^) during a series of cruises conducted during 1987-89 ranged from 68 to 203 mm TL (mean size= 154.5 mm TL; Fig. 2C ); and most fishes li •' VIMS, Department of Fisheries, Gloucester Point, VA 23062. Unpubl. data. Table 4 Likelihood ratio tests comparing von Berta plagiusa, from Chesapeake Bay, Virginia. an df ffy parameter estimates for = degrees of freedom. male 1 1 1 and female ( 2 ) blackcheek tonguefish Symphurus Hypothesis Linear constraints Residual SS x\ df P HU none 99,760.5 Ho)l ^.,=^.2 99,780.9 0.1043 1 0.7467 Ho)2 K, =K, 99,765.8 0.0271 1 0.8692 Ho).3 'oi ~ '02 99,796.9 0.1861 1 0.6661 Hco4 ^.1=^«. if, = K.^ 'oi ~ 'o2 10,051.6 6.558 3 0.0874 Terwilliger and Munroe: Age, growth, longevity, and mortality of Symphurus plagiusa 351 B. 0 c 0) measured 140-180 mm TL. Lengths of fishes collected within the Bay sys- tem completely overlap those taken from the nearby shelf. Munroe ( 1998 ) noted that most blackcheek tongue- fishes in collections from throughout the species' geographic range were usually much smaller than the larg- est sizes known for the species be- cause only 4% of 568 fish he exam- ined exceeded 164 mm TL. Wenner and Sedberry (1989) sampled 8780 black- cheek tonguefishes from coastal habi- tats ( < 10 m depth ) between Cape Fear, North Carolina, and the St. Johns River, Florida, and reported that length composition of trawl-caught S. plagiusa was consistent from season to season, with mean sizes ranging between 140-150 mm TL. The larg- est specimens taken during that study were 210 mm TL, and few fish <100 mm TL were caught, possibly because fish smaller than this inhab- ited a different habitat as juveniles or because they did not recruit to the fish- ing gear until 100 mm TL. On the con- tinental shelf off Tampa Bay, Florida, lengths (124-174 mm TL) for black- cheek tonguefishes were also similar to those reported for fish from Chesa- peake Bay (Moe and Martin, 1965). Although age estimates are not available for other populations of blackcheek tonguefish, a general comparison of growth between black- cheek tonguefishes occurring in Chesapeake Bay with those from other areas is provided by examination of length- weight parameters derived for different populations. Dawson (1965) generated a length-weight relation- ship for 3504 blackcheek tonguefish (43-148 mm TL) occurring in Gulf of Mexico waters. With this rela- tionship, weights were generated from 510 randomly selected lengths between 43 and 148 mm TL and these were then compared with a similar number of weights generated for fishes of the same size range by using the length-weight relationship derived for blackcheek tonguefishes from Chesapeake Bay. No significant differences (^=0.53; df=1010;P=0.60) were evident in this comparison between weights of fishes of these two regions, indicating that growth in weight, at least, is similar for fishes of this size range occur- ring in these widely separated areas within the spe- cies' range. n = 172 200 - [ A 180- « 160 - ■ ♦ L ^ • 140 -^ ■ ♦ -r * 120- „.--- •" 1 100 • j ^ ^^ 80- Y 60 ■ i 40 ■ 20- 0 • — 1 — — 1 — 1 — 1 — — 1 200 - 180 • B n =275 160 ■ 140 ■ 1 120 • » X-- X 100 • ■ ^ ^\ * 80 • 60- 40 ■ • > 20 ■ n • ^ 1 1 — 1 1 Age (years) Figure 9 Plot of back-calculated and predicted lengths from the von Bertalanffy growth model for blackcheek tonguefish, Symphurus plagiusa, from lower Chesapeake Bay and major lower tributaries: (A) males; (B) females. Annuli form on the otoliths of blackcheek tongue- fishes from late spring through early summer (mostly June), coincident with the summer warming period and also at the initiation of the spawning season (Terwilliger, 1996). For C. abbreviatus, a tonguefish occurring in temperate seas off eastern China (Zhu and Ma, 1992), annuli form on otoliths during March- May, a period also corresponding to seasonal timing of gonad maturation and warming of water tempera- tures. Koski (1978) reported that annuli form on scales of hogchokers in the Hudson River from April to mid-June, which is a period of increasing water temperature, and for mature fish this also corre- sponds to the time just prior to and overlapping the early spawning season. Flatfishes typically exhibit sexual dimorphism in size at age (Roff, 1982; and others, see below) — 352 Fishery Bulletin 97(2), 1999 females usually growing larger than males. In the Chesapeake Bay population, male and female blackcheek tonguefishes attain similar maximum sizes (mean value for oldest females=181.2 mm TL; mean value for oldest males=177.8 mm TL). In sum- marizing size information for blackcheek tongue- fishes collected throughout the geographic range of the species, Munroe (1998) also reported no signifi- cant size differences between males (to 174 mm TL) and females (to 172 mm TL). For 23 other species of western Atlantic tonguefishes, females reached larger sizes than males in 11 of these species, in six others a larger size was reported for males, whereas in seven other species similar maximum sizes were attained by both sexes (Munroe, 1998). However, since no age estimates have been made for these other species of Syryiphurus. comparative data on dimor- phism in lengths at age for these symphurine tonguefishes are unavailable. For other cynoglossids with accompanying age data, considerable variation exists in the degree of sexual dimorphism evident in sizes attained by the sexes. In the Malabar sole iCynoglossus serriifasciatus) off India, females grow larger than males, although reported differences in maximum sizes between sexes were relatively small (Seshappa and Bhimacher, 1954). Similarly, although female C. abbreviatus ( to 348 mm TL) from Jiaozhua Bay, China, were larger than males (to 321 mm TL), average total lengths at age between the sexes were not significantly different (Zhu and Ma, 1992). Fe- males of C. biUneatiis and C. arel were also reported to be larger than males sampled from the same popu- lations (Hoda and Khalil, 1995), whereas sizes for adult female (83-173 mm TL) and male (84-168 mm TL) C. macrostomus were found to be similar (Victor, 1981). Female C. canariensis also grow larger than males (Chauvet, 1972). For C. semilaevis inhabiting the Bohai Sea, China (Meng and Ren, 1988), quite a different situation exists. In this species, females at- tain lengths (to 638 mm TL) slightly more than twice those reached by males (to 312 mm TL). In other es- tuarine flatfish, such as the hogchoker, which occurs throughout much of the geographic range of the blackcheek tonguefish, females grow to larger sizes than do males from the same population (Koski, 1978). For most species of the Pleuronectidae, females also typically grow larger and have both a later age and larger size at maturity than do males (Cooper^). Back-calculated lengths-at-age for blackcheek tonguefishes were consistently less than observed lengths-at-age for all age groups combined, which may be attributed to Lee's phenomenon. The princi- ■* Cooper, J. A. 1998. National Marine Fisheries Service, Na- tional Systematics Laboratory, Wa.shington, DC 20.560. Per- sonal common. pal cause of Lee's phenomenon in unexploited fish populations, such as that of blackcheek tonguefish inhabiting Chesapeake Bay, is that faster growing fish tend to mature and die earlier than do smaller members of the same year class (Gerking, 1957; Ricker, 1975). When a larger proportion of older fish die, the result is a smaller estimated size for fish at younger ages than the true average size at the age in question. Since blackcheek tonguefishes are not harvested commercially or recreationally within the Bay, the presence of Lee's phenomenon in the back- calculation estimates may be due to higher mortal- ity of larger individuals within an age class, or pos- sibly this finding may reflect the fact that larger fish move out of the Bay system and onto the nearby shelf region. The ages of blackcheek tonguefishes occur- ring on the continental shelf throughout the species' range are unknown. No offshore samples of black- cheek tonguefishes were collected during this study, and it is unknown if any of these fish are older than those occurring in Chesapeake Bay. Although female blackcheek tonguefishes captured in Chesapeake Bay gi-ew at a slightly faster rate than did males, observed differences in back-calculated lengths between male and female blackcheek tonguefishes beyond age 4 were not statistically sig- nificantly different. However, our sample sizes for older fish were small and interpretation of data for this oldest age class is limited. In other cynoglossids, such as Malabar sole, faster growth has also been observed in females compared with that for males (Seshappa and Bhimachar, 1954). Rajaguru (1992) found that female C. arel grew faster than males, but no significant differences in growth patterns were found between female and male C. lida. After their first year, female hogchokers in the Hudson River also grew faster than males (Koski, 1978). In some species of the Soleidae, such as the Agulhas sole (Austroglossus pectoralis) occurring off South Africa (Zoutendyk, 1974) and Solea solea in Spanish wa- ters (Ramos, 1982), females also have a faster growth rate than do males within the same populations. A faster growth rate for females compared with that of males has also been reported in a variety of other flatfish species (Chen et al., 1992; Santos, 1994). Estimates of the growth parameter /C( Table 5) for blackcheek tonguefish are relatively high, indicat- ing that these fish reach their asymptotic length rela- tively rapidly (Francis, 1996). Few estimates of K- values are available for other tonguefishes, but for those species studied, growth rates are relatively high. Where data are available, all, except two spe- cies, have higher growth coefficients reported for fe- males. In r. arel, AT-values were 0.315 for females and 0.238 for males, and in C. lidci A'-values of 0.233 Terwilliger and Munroe: Age, growth, longevity, and mortality of Symphurus plagiusa 353 for females and 0.223 for males were reported (Rajaguru, 1992). For C. abbreviatus (Zhu and Ma, 1992), /C- values were 0.395 for fe- males and 0.344 for males. In two of the largest species in the fam- ily, C. semilaevis and C. canar- iensis. the estimated /C-values were higher for males compared with those for females. In C. semilaevis, K-value for females was 0.264, and that of males was 0.352 (Meng and Ren, 1988). For female C. senegalensis the K- value was 0.32 compared with 0.36 for males (Chauvet, 1972). In Chesapeake Bay, blackcheek tonguefishes achieve most of their growth in length during their first year. Males and fe- males attained up to SO^f of their total growth in length by the end of their first year, and up to 72% of their total growth by the end of their second year. Such rapid first-year growth does not appear to be unusual among members of the Cynoglossidae. For example, male C. arel grew between 180-194 mm TL, and females from 201-210 mm TL, during their first year (Rajaguru, 1992), which represented up to 58% and 51% of their total growth in length (333 and 393 mm, respectively). Similar to the pattern observed in blackcheek tonguefish, second-year growth in this species was estimated to be only 18% and 21% of the total growth, respectively, for males and females. In C. lida, males (151-154 mm TL) and females (153-156 mm TL) also reached fairly large sizes during their first year of life, with first-year growth for males and females ranging between 69 and 71% of the total growth in length (216 and 218 mm, respectively), compared with second-year growth values of only 13-14% of the total length (Rajaguru, 19921. Cynoglossus macrolepidotus also has a high growth rate during its first year, reach- ing a length of about 160 mm TL, and like the pre- ceding species, during the second and subsequent years there occurs a marked reduction (only 10-30 mm increase/yr) in growth rate (Kutty, 1967). Like- wise, in C. abbreviatus (Zhu and Ma, 1992), first- year growth is maximal for both sexes (males to 162 mm TL, females to 178 mm TL; representing 50% and 51% of total growth, respectively), whereas sec- ond-year growth in this species averages only about 16% or less of the total growth. This growth pattern was also apparent for C. canariensis where male and Table 5 Comparision of von Bertalanffy parameters of selected Pleuronectifornies occuring in temperate estuaries of the western North Atlantic. Species ^arameters Source Paralichthys dentatus ( i o ) K: 0.215 Desfosse, 1995 L„ 859 mm to- -0.690 Paralichthys lethostigma ( + ) K. 0.235 Miller etal., 1991 L^ 760 mm to- 0.570 Scophthalmus aquosus ( S ) K: 0.272 Miller et al., 1991 ^» 383 mm 'o: 0.418 Pseudopleuronectes ameriranus ( 2 ) K-. 0.217 Miller etal., 1991 L^ 375.8 mm to- 0.730 Symphurus plagiusa ( '+ o ) K-. 0.308 Terwilliger, 1996 L^ 192.4 mm to- -0.832 Trinectes maculalus ( S ) K-. 0.195 Miller etal., 1991 L„ 209.5 mm to- -0.353 female fish, respectively, achieved 50%c and 55%f of their total growth in length during the first year, compared with second-year growth equalling only 14-15% of the total growth in length (Chauvet, 1972). In another large-size species, C. semilaevis, first-year growth is also rapid, with males reaching 141 mm TL at end of year one and females growing to 194 mm TL, values that represent 45% and 30% of the total growth in length achieved by this species (Meng and Ren, 1988). However, in contrast to other cynoglossid species examined thus far, growth in this species during the second year differs markedly be- tween the sexes. For females, the average growth increment during the second year was greater than that of the first year (to 307 mm TL, 36% of total growth in length), whereas during this same time interval males averaged only about one-half of the growth increment attained during year one (to 204 mm TL, representing 20% of total growth). In the hogchoker, growth in length is also rapid; males and females achieve 56% and 44% of their total growth in length during the first year (Koski, 1978). By the end of their second year, hogchokers have reached on average 74% (males) and 63% (females) of their total growth in length. The substantial reduction in growth rate beyond age 1 noted for blackcheek tonguefishes occurs at a time coincident with maturation (Terwilliger, 1996). Roff (1982) noted that, for fish in general, it is fre- quently observed that rate of growth decreases with 354 Fishery Bulletin 97(2), 1999 the onset of maturity. This slowing in somatic growth may reflect an allocation of energy to gonadal devel- opment, or may be a response to other physiological processes associated with maturation. Other cynoglossid species that have been aged also mature sexually at a relatively early age even in some of the larger-size species. Seshappa and Bhimachar ( 1954) reported that young Malabar soles (C. seryiifasciatus ) grew to adult size within a year or less and that af- ter maturation, growth slowed distinctly. After an initially high growth rate during their first year, there also occurred a marked reduction in the growth rate of C. macrolepidotus during their second year, at a time coincident with initiation of the spawning season (Kutty, 1967). Likewise, in other cynoglossids (C. arel and C. lida) occurring off India, growth dur- ing the first year is also rapid and this fast growth period is then followed by a considerable reduction in the growth rate during the next year when the fishes matured sexually (Rajaguru, 1992). In C arel, for example, which reaches total lengths to 333 mm SL, males and females mature at sizes of about 217 mm TL and 225 mm TL, respectively, sizes reached by this species at the beginning of their second year of life. In C. lida, size at 50^?^ maturity is 167 mm TL for males, and 179 mm TL for females. These repre- sent sizes that are reached by the fish during the beginning of their second year of life. For C. dubiiis (Seshappa, 19761 minimum size at maturity was re- ported to be about 287 mm TL for females, a size that corresponded to fish approximately 2-3 years old. In C. canariensis, a species reaching total lengths of 486 mm (males) and 519 mm (females), maturity occurs at about 1.5 years when fish are about 300 mmTL(Chauvet, 1972). Hoenig's and Royce's mortality estimates, based on the maximum known age of a species, indicate a relatively high instantaneous total mortality in the blackcheek tonguefish population inhabiting Chesa- peake Bay (Table 6). Relatively few of the blackcheek tonguefishes within Chesapeake Bay are older than three years. The age structure observed in this popu- lation may, however, reflect the relatively short life span of this species (5-i- years), or it could also reflect high mortality levels experienced by blackcheek tonguefishes in the Bay environment; undoubtedly it would also be influenced by the emigration of larger individuals from the Bay system. Relatively short life spans may be typical of spe- cies of Symphiirus and of the Cynoglossidae, in gen- eral. In fact, some dwarf species of tropical Symphurus reach maximum sizes of only 35—45 mm SL (Munroe, 1990, 1998) and probably live no more than a year. Age estimates based on growth marks interpreted from scales of tonguefishes occurring in coastal waters off India indicate that Malabar sole reach an age of only about 2+ years (Seshappa and Bhimachar, 1954). Based on studies of tonguefishes from the west coast of India, a life span of 3-4 years was reported by Seshappa (1978, 1981) for C. lida (size to 220 mm TL; most 140-189 mm TL) and C. puncticeps (to 209 mm TL; most 90-179 mm TL), whereas a life span of 6-i- years (to 339 mm TL; most 160-299 mm TL) was reported for C. bilineatus from the same region. More recently, Rajaguru (1992) re- ported that C. arel and C. lida along the southeast coast of India have life spans just over 3 years. For C. dubius off India (to 414 mm TL), age estimates were 6-i- years for most individuals, although some individuals reached 10 years (Seshappa, 1976). Cyno- glossus arel off Taiwan (Lin, 1982) live to be at least 4 years. Kutty (1967) reported a maximum age of 6-7 years for C. macrolepidotus (to 330 mm TL). For C canariensis (to 519 mm TL) off tropical West Af- rica, both sexes live to at least 8 years (Chauvet, 1972). For the temperate species, C abbreviatus, from Jiaozhua Bay, China ( Zhu and Ma, 1992 ), which is commercially exploited, females reach 8 years and males live to 7 years, and total mortality for this population was estimated to be very high (0.607 ). For Table 6 Mortalitv estimates (2) for various northwestern Atlantic Pleurunectiformes ba sed on Hoenig's and Royce's equations. Values in parentheses reflect extrapolated mortality rates for a possible seventh age class of blackcheek tonguefish, Symphurus plagiusa, \ found in lower Chesapeake Bay and the inner continental shelf. Species Maximum age Hoenig's equation Royce's equation Data source Winter flounder iP. americanus) 14 years 0.32 0.33 Lux, 1973 Summer flounder (P dcnto/wsl 9 years 0.49 0.51 Desfosse, 1995 Windowpane (S. a<7U0.su.sl 7 years 0.62 0.66 Moore, 1947 Southern flounder i.P. lethostigma) 6 years 0.7.3 0.77 Music and Pafford, 1984 Hogchoker (T! maculatus) 6 years 0.73 0.77 Mansuetti and Pauly, 1956 Blackcheek tonguefish (S. plagiusa 1 5 years (6 years) 0.87 (0.73) 0.92(0.77) Terwilliger. 1996 Terwilliger and Munroe: Age, growth, longevity, and mortality of Symphurus plagiusa 355 C. semilaevis from Chinese waters, a strong sexual dichotomy in longevity was apparent in the popula- tion studied; females reached 14 years (most 2-8 years), whereas males in this population attained a maximum age of only 5 years (Meng and Ren, 1988). By comparison, longevity estimates for a variety of other flatfishes occurring in temperate latitudes and from families other than the Cynoglossidae range from 6 to 30 years (Devoid, 1942; Arora, 1951; Pitt, 1967; Lux and Nichy, 1969; Lux, 1970. 1973; Koski, 1978; Wada, 1970; Zoutendyk, 1974; Smith and Daiber, 1977). As stated earlier, the population of S. plagiusa in Chesapeake Bay represents the northernmost loca- tion of a major population of this species (Munroe, 1998). In the western North Atlantic, variability in abiotic conditions increases from southern to north- ern latitudes (Parr, 1933; Miller et al., 1991). Abiotic variability also increases in environments located closer to the shoreline (Russell-Hunter, 1970; Miller et al., 1991 ). Fluctuations in abiotic factors, especially seasonal extremes of temperature during winter, may contribute to the high level of total instantaneous mortality measured for blackcheek tonguefishes within Chesapeake Bay. Bottom trawls conducted in this system during late winter and early spring of some years ( 1996, for example ) catch numerous dead and moribund tonguefishes (Geer'^), apparently due to stresses associated with extremely cold water tem- peratures (ca. 2°C. in 1996). More extreme winter temperatures and longer duration of cold tempera- tures may be a factor limiting the establishment of populations of blackcheek tonguefishes in more northern estuaries, where this species is known only by the occurrence of newly settled juveniles. In coastal estuaries throughout its U.S. geographic range, blackcheek tonguefishes appear to be sensitive to ex- treme winter temperatures, because winter mortali- ties have also been reported for this species in estu- aries even as far south as those in Texas (McEach- ron et al., 1994). In addition to causing death out- right, stresses associated with extreme winter tempera- tures may also render weakened tonguefishes easier prey to predators and thus further contribute to win- ter mortalities. Mortality estimates for the population of black- cheek tonguefishes residing in Chesapeake Bay may also be inflated due to emigration of larger individu- als out of the Bay and onto the nearby continental shelf Hildebrand and Cable ( 1930) earlier had sug- gested that adult blackcheek tonguefishes in North Carolina estuaries undergo a seaward migration during May-August. An autumn emigration of blackcheek tonguefishes from Chesapeake Bay to the nearby shelf region just outside of the Bay may be reflected in the significantly larger catches (mean no./tow) of blackcheek tonguefishes reported (VIMS*') in this region during fall sampling ( 1987, 75.4/tow; 1988, 25.7/tow) compared with springtime catches (1987, 11.0/tow; 1988, 4.37/tow; 1989, 2.2/tow). Data supplied in Wenner and Sedberry (1989) also sup- port the hypothesis that there is a seasonal migra- tion of larger tonguefishes out of estuaries and onto the coastal shelf off South Carolina. They reported that this species was widespread throughout the survey area but noted that significant seasonal dif- ferences were apparent in the frequency of occurrence and abundance of S. plagiusa in coastal habitats. This species occurred less frequently during winter (48*^ of all tows; 566 individuals) than in spring (85% of tows; 1,460 individuals), summer (76% of tows; 798 individuals), or fall (94% of tows; 5962 individu- als). Catches (mean no./tow) of this species were low- est in summer (8/tow) and winter (9/tow), higher in spring (2 1/tow), and highest during the fall (62/tow), when the largest fish (up to 210 mm TL) were taken. Although Wenner and Sedberry (and other studies) have reported that it is usually only larger-size blackcheek tonguefishes that are captured on the inner shelf region, no studies have verified any sea- sonal or spawning movements of blackcheek tonguefishes out of the estuaries, or the amount of exchange of individuals within populations between estuarine embayments and adjacent neritic habitats on the inner shelf It is also unknown whether all larger individuals undergo an ontogenetic migration out of estuaries onto the nearby shelf; what portion of a population residing in estuaries moves out onto the shelf at any given age; what the estuarine resi- dency time is for larger tonguefishes, especially in such larger and deeper estuaries as the Chesapeake Bay which may be utilized by larger tonguefishes for longer periods of time; or even if further exchange of individuals between these two environments occurs once an individual has emigrated beyond the estua- rine boundary. Analysis of catch curve data (Fig. 10) may also lend insight on inflated mortality estimates. Although the bimodality of the catch curve data in- fers possible variability in year-class strength that invalidates its use for mortality estimations ( Robson and Chapman, 1961; Everhart and Youngs, 1981), the data suggest that a seventh year class may ex- ist, possibly on the nearby shelf region. If so, mortal- ity estimates based on a maximum age of 5+ years ^Geer.P. J. 1996. VIMS, Department of Fisheries, Gloucester Point, VA 23062. Personal commun. ^ VIMS, Department of Fisheries, Gloucester Point, VA 23062. Unpubl. data. 356 would be inflated compared with those calculated for an older year class. Sex ratio of tonguefishes sampled during this study devi- ated from 1:1 mostly due to large single-sex catches of females dur- ing the summer months in the lower Bay. In lower Chesapeake Bay, apparently some spatial separation of the sexes occurs (Terwilliger, 1996). Here, dense concentrations of female black- cheek tonguefishes were com- monly found in the deeper areas, and fewer males were taken in the same trawls. Among sympa- tric species of pleuronectiforms, the hogchoker also reportedly displays bathymetric separation of the sexes in tributaries of Chesapeake Bay during the sum- mer months prior to spawning (Mansuetti and Pauly, 1956), when females occupy grass beds and shoal regions, while males are surmised to occupy deeper ar- eas withm the estuary. This pattern of habitat seg- regation by adult hogchokers, however, was not evi- dent in the study conducted by Koski (1978) in the Hudson River, although he noted that although the overall sex ratio for fishes examined did not differ from 1;1, some individual trawl catches were domi- nated by or consisted entirely of one sex. Significant differences in the sex ratios of catches of other tonguefishes have also been reported. Seshappa and Bhimachar ( 1955) noted that sex ratios of catches of Malabar sole during the spawning period were mark- edly different than those made during other times of the year and attributed these differences to differ- ential behavior of the two sexes during their spawn- ing migrations. Hoda and Khalil ( 1995) also reported skewed sex ratios (1.0:2.55 males to females) for catches of^ Cynoglossus arel, whereas those (1.16:1) for C. bilineatus were not significantly different from 1:1. A skewed ratio (1.56:1) favoring females over males was also found for C. semilaevis (Meng and Ren, 1988). Although some studies have inferred spatial differences in movements of the different sexes of cynoglossid tonguefishes based on fisheries catch data, it is difficult to draw any general conclu- sions from these data. Just as we lack reliable docu- mentation concerning spatial movements for the two sexes of blackcheek tonguefish, directed studies pro- viding the detailed information necessary for inter- Figure 10 Catch curve for blackcheek tonguefish, Symphurus plagiusa. collected by otter trawl in lower Chesapeake Bay and major lower tributaries, 1993-95. Numbers above plot represent numbers at each age. preting adult movements of any of these other cynoglossid species have yet to be performed. North temperate estuaries, such as Chesapeake Bay, are rigorous physical systems well known for environmental extremes in temperature, salinity, and dissolved oxygen, and for strong seasonal cycles of primary and secondary production. However, the extensive distribution within the Bay and its tribu- taries of shallow, soft bottom sediments presents con- siderable amounts of suitable habitat for flatfishes. Twelve species of flatfishes have been recorded from the Bay (Murdy et al., 1997). The unique geographi- cal location of Chesapeake Bay at the northern end of a warm temperate and southern end of a cold tem- perate region (Briggs, 1974) may account for the oc- currence within the Bay of diverse faunal compo- nents, which have either warm or cold water affini- ties. Among pleuronectiforms inhabiting Chesapeake Bay, for example, are six species representing five different families (Cynoglossidae, Achiridae, Para- lichthyidae, Scophthalmidae, and Pleuronectidae). This assemblage includes representatives of families whose species occur primarily in cold temperate or boreal areas (Scophthalmidae, Pleuronectidae) or in warm temperate and tropical regions (Cyno- glossidae, Achiridae, Paralichthyidae). Of the spe- cies occurring within Chesapeake Bay are blackcheek tonguefish, hogchoker, and smallmouth fiounder. Terwilliger and Munroe: Age, growth, longevity, and mortality of Symphurus plogiusa 357 Etropus microstomus (to ca. 159 mm TL), small-size species that are not commercially or recreationally exploited and ones that complete their life cycles within the Bay. Virtually nothing is known concern- ing age and growth of smallmouth flounder, espe- cially within the Bay system; therefore further com- parisons of its life history with those of sympatric flatfishes are limited. The flatfish fauna of the Bay also includes summer flounder [Paralichthys dentatus), southern flounder (Paralichthys lethostigma ), window- pane (Scophthalmus aquosiis), and winter flounder (Pseudopleuronectesamericanus), species that do not necessarily complete their life cycles entirely within Chesapeake Bay and ones that are heavily exploited both commercially and recreationally. The remain- ing species of flatfishes recorded from the Bay are transient species that occur only irregularly within the Bay and will not be discussed further. The diver- sity in size and growth parameters represented among those flatfishes occurring in Chesapeake Bay which have been studied provides an interesting opportunity for investigating comparative life histo- ries of sympatric members of an estuarine benthic fish assemblage with somewhat similar lifestyles (Table 5). Few published age-growth studies exist for noncommercially exploited flatfishes, especially for those species occurring in temperate estuaries of the western North Atlantic. Von Bertalanffv growth pa- rameters for species other than blackcheek tongue- fish were taken from data found in the literature ! (Miller et al., 1991; Desfosse, 1995). These data re- ' veal that among this assemblage, blackcheek tongue- fish have the highest value for growth parameter K (Table 51, and therefore reach asymptotic length faster than sympatric counterparts (Francis, 1996). Of sympatric flatfishes occurring in Chesapeake Bay, only hogchokers display growth parameters and maturation schedules (some males mature at end of first year, most of both sexes mature by age 2; Koski, 1978) similar to those observed for blackcheek tonguefish. In contrast, summer and southern floun- ders mature at ages 2-3, windowpane flounders mature at ages 3 or 4, and winter flounders mature at ages 2-5. Later age at maturation in these spe- cies reflects a different life history pattern, and one that allows for energy allocation favoring increased somatic growth during several years prior to the on- set of maturation. Because these other species ma- ture at later ages than do blackcheek tonguefish and hogchoker, growth in length for these fishes would be relatively rapid over the course of several years during the early part of their lives (Fig. 11). Estimates of asymptotic length, reflecting average maximum length attained by a species, also varies considerably among members of this assemblage. Asymptotic lengths for summer flounder (859 mm TL ) and southern flounder ( to 760 mm ) greatly exceed those for blackcheek tonguefish and hogchoker ( 192 and 210 mm, respectively). In fact, these fishes grow more during their first year than do blackcheek tonguefish and hogchokers in a lifetime. Windowpane and winter flounder (383 and 376 mm, respectively) attain asymp- totic lengths intermediate to these other two groups of sympatrically occurring estuarine flatfishes. Estimates of instantaneous total mortality based on blackcheek tonguefishes caught within Chesa- peake Bay during this study are also relatively higher when compared with those for most other estuarine dependent species of northwestern Atlantic Pleuro- nectiformes (Table 6). In this comparison, blackcheek tonguefish are among the shortest-lived species and therefore have one of the higher rates of instanta- neous total mortality. Relatively few individuals within Chesapeake Bay are older than three years. Extrapolating mortality rates for a possible seventh year class (age 6+ ), which may be found on the inner shelf region, provides an instantaneous total mor- tality estimate comparable to those for hogchokers and southern flounder caught in Chesapeake Bay. Blackcheek tonguefish exhibit the youngest age at first maturity, the highest growth coefficient, the youngest known maximum age, and the smallest asymptotic length when compared with sympatric pleuronectiforms from Chesapeake Bay. Blackcheek tonguefish exhibit rapid growth until maturity, af- ter which growth rate declines markedly. Munroe (1990, 1998) indicated that eight dwarf species of Symphurus reach sexual maturity between 28 and 45 mm SL. Small size and presumably early age at maturation may be characteristic of species in this genus (Munroe, 1988); even S. jenynsi, the largest species in the genus (reaching ca. 350 mm TL), ma- tures at a relatively small size (about 132 mm TL). More ageing studies on fishes of this genus are re- quired to test this hypothesis. Other cynoglossid flat- fishes, although they attain a relatively large size before reaching sexual maturity, are only 1 or 2 yr old when first maturing. According to available data, it is quite possible then that tonguefishes in general have an age dependent maturity schedule. Roff ( 1982) indicated that age, and not size, would appear to be most important in fish species that mature early. Data for tonguefishes support that finding. When compared with estimates for sympatric flat- fishes, blackcheek tonguefish caught during this study exhibit the highest rates of instantaneous to- tal mortality. Other species inhabiting the Bay sys- tem also mature at larger sizes, have larger asymp- totic lengths, and slower growth rates. Interesting 358 Fishery Bulletin 97(2), 1999 2 3 4 5 6 7 Age Figure 11 Von Bertalanffy growth curves for selected Pleuronectiformes occurring in western North Atlantic temperate estuaries. Values for species other than Symphurus plagiusa were derived from the literature. to note, it is only the smallest-size and shortest-lived flatfishes (S. plagiusa, T. macidatus, and probably E. microstomus ) that complete their life cycles within Chesapeake Bay. Under an environmental regime with a large component of unpredictable, nonselec- tive mortality, an organism is hypothesized to allo- cate a larger portion of its resources to reproductive activities (Adams, 1980). Life history parameters, such as rapid growth, small body size, early age at maturity and relatively short life span, characteris- tics featured in these small-size flatfishes complet- ing their life cycles within Chesapeake Bay, are con- sistent with those observed for other fish species that survive in and successfully exploit seasonally dy- namic and highly unpredictable environments. Acknowledgments This study represents part of a M.A. thesis (by Mark R. Terwilliger) at the School of Marine Science, Vir- ginia Institute of Marine Science, College of William and Mary. We thank Joy Dameron, Deane Estes, Eric Farrar, Pat Geer, Paul Gerdes, Dave Hata, Mike Land, Todd Mathes, Dee Seaver, Mike Seebo, and Chandell Terwilliger of VIMS for assisting in field collections and for providing data and specimens. We extend our appreciation to Herb Austin for allowing us access to sources of unpublished data archived in the VIMS Fisheries Department. This study bene- fited from financial support arranged by Herb Aus- tin, Jack Musick, Gene Burreson, and Joe Loesch (VIMS). Additional financial support was provided by the VIMS minigrant committee, the student con- ference fund of the College of William and Mary, and by the American Institute of Fisheries Research Bi- ologists' student conference fund. Chandell Terwil- liger provided moral support throughout the course of the study. Pat Geer, Dave Hata, Steve Bobko (ORST), and Mike Murphy (FL DEP) provided valu- able insight on statistical analyses. A special thanks to Pat Geer for his help with trawl survey data analy- sis and manipulation. Tom Sminkey (FL DEP) pro- vided useful instruction on use of the Biosonics Op- tical Pattern Recognition System, and Dave Harshany (FL DEP) helped with image production Terwilliger and Munroe: Age, growth, longevity, and mortality of Symphurus plagiusa 359 using Optimas software. Martha Nizinski (VIMS) assisted with figure preparation. Sandra Raredon, Division of Fisheries, Smithsonian Institution, pro- vided assistance with manuscript translation. Roy Crabtree, Tom Sminkey (FL DEP), Kathy Lang (NMFS, Woods Hole), Bruce Collette (NMFS System- atics Laboratory), and three anonymous reviewers critically reviewed earlier drafts of the manuscript. Literature cited Adams, P. B. 1980. Life history patterns in marine fishes and their con- sequences for fisheries management. Fish. Bull. 78:1-12. Arora, H. L. 1951. An investigation of the California sand dab, Cithar- ichthys sordidus (Girard). Calif Fish Game 37(l):3-42. Bagenal, T. B., and F. W. Tesch. 1978. Age and growth, /fj T. B. Bagenal (ed.). Methods for assessment offish production in fresh waters, 3rd ed., p. 101-136. Blackwell Scientific Publications, Oxford. Barbieri, L. R. 1993. Life history, population dynamics and yield-per-re- cruit modeling of Atlantic croaker, Micropogonias undu- latus. in the Chesapeake Bay area. LInpubl. Ph.D. diss., College of William and Mary, Williamsburg, VA, 140 p. Biosonics, Inc. 1987. Optical Pattern Recognition System data acquisition program manual, version 1.08. Biosonics, Inc., Seattle, WA, 112 p. Bonzek, C. F., P. J. Geer, J. A. Colvocoresses, and R. E. Harris Jr. 1993. Juvenile finfish and blue crab stock assessment pro- gram, bottom trawl survey, annual data summary. An- nual data summary report series, vol. 1992. Special Scien- tific Report 124, Virginia Institute of Marine Science, Col- lege of William and Mary, Gloucester Point, VA, 217 p. Briggs, J. C. 1974. Marine zoogeography. McGraw-Hill Book Co., New York, NY, 475 p. Cerrato, R. M. 1990. Interpretable statistical tests for growth comparisons using parameters in the von Bertalanffy equation. Can. J. Fish. Aquat. Sci. 47:1416-1426. Chauvet, C. 1972. Croissance et determination de I'age par lecture d'ecailles d'un poisson plat de Cote d'lvoire Cynoglossus cananensis (Steind. 1882). Doc. Scient. Cent. Rech. Oceanogr. Abidjan 3:1-18. Chen, D., C. Liu, and S. Dao. 1992. The biology of flatfish (Pleuronectidae) in the coastal waters of China. Neth. J, Sea. Res. 29:25-33. Chittenden, M. E., Jr., and S. D. McEachran. 1976. Composition, ecology, and dynamics of demersal fish communities on the northwestern Gulf of Mexico continen- tal shelf, with a similar synopsis for the entire Gulf Texas A&M University TAMU-SG-76-208, 104 p. Dawson, C. E. 1965. Length-weight relationships of some Gulf of Mexico fishes. Trans. Am. Fish. Soc. 94(3):279-280. Desfosse, J. C. 1995. Movements and ecology of summer flounder, Para- tichthys dentatus, tagged in the southern mid-Atlantic bight. Unpubl. Ph.D. diss., College of William and Mary, Williamsburg. VA, 187 p. Devoid, F. 1942. Plaice investigations in Norwegian waters. Rep. Norw. Fish. Man Invest. VIIlSl, 83 p. Everhart, W. H., and W. D. Youngs. 1981. Principles of fishery science, 2nd ed. Cornell Univ. Press, Ithaca, New York, NY, 349 p. Francis, R. L C. C. 1996. Do herring grow faster than orange roughy? Fish. Bull. 94:783-786. Geer, P. J., H. M. Austin, and C. F. Bonzek. 1997. Juvenile fish and blue crab stock assessment program bottom trawl survey annual data summary report. An- nual data summary report series, vol. 1996. Special Scien- tific Report 124, Virginia Institute of Marine Science, Col- lege of William and Mary. Gloucester Point, VA, 275 p. Geer, P. J., J. A. Colvocoresses, H. M. Austin, and C. Bonzek. 1993. Estimation of relative abundance of recreationally important finfish in the Virginia portion of Chesapeake Bay: annual progress report. U.S. Fish and Wildlife Ser- vice Sportfish Restoration Project F104R2, College of Wil- liam and Mary, Virginia Institute of Marine Science, Gloucester Point, VA, 105 p. Gerking, S. D. 1957. Evidence of ageing in natural populations of fishes. Gerontologia 1:287-305. Ginsburg, I. 1951. Western Atlantic tonguefishes with descriptions of six new species. Zoologica (NY) 36:185-201. Heidinger, R. C, and K. Clodfelter. 1987. Validity of the otolith for determining age and growth of the walleye, striped bass, and smallmouth bass in power plant cooling ponds. In R. C. Summerfelt and G. E. Hall (eds. ), Age and growth in fish, p. 241-251. Iowa State Univ. Press, Ames, lA. Hildebrand, S. F., and L. E. Cable. 1930. Development and life history of fourteen teleostean fishes at Beaufort, North Carolina. Bull. U.S. Bur Fish. 43:383-488. Hildebrand, S. F, and W. C. Schroeder. 1928. Fishes of Chesapeake Bay Bull. U.S. Bur. Fish. 43:1-366. Hoda, S. M. S., and B. Khalil. 1995. Observations on the biology of Cynoglossus bilineatus (Lacep.) and C. Arel (Bl. & Schn.) from the Karachi-Sindh coast. In M. F. Thompson and N. M. Tirmizi (eds.). The Arabian Sea: living marine resources and the environment, p. 309-330. Vanguard Books Ltd., Lahore, Pakistan. Hoenig, J. M. 1983. Empirical use of longevity data to estimate mortal- ity rates. Fish. Bull. 82:898-902. Jearld, A., Jr. 1983. Age determination. In L. A. Nielsen and D. L. Johnson (eds.). Fisheries techniques, p. 301-324. Am. Fish. Soc, Bethesda, MD. Jorgenson, S. C, and G. L. Miller. 1968. Length relations of some marine fishes from coastal Georgia. U.S. Fish Wildl. Serv. SSR-Fisheries, No. 575, 16 p. Kimura, D. K. 1980. Likelihood methods for the von Bertalanffy growth curve. Fish. Bull. 77:765-775. Koski, R. T. 1978. Age, growth, and maturity of the hogchoker, Trinectes maculatus, in the Hudson River, New York. Trans. Am. Fish. Soc. 107:449-453. 360 Fishery Bulletin 97(2), 1999 Kutty, M. K. 1967. Observations on the growth and mortahty of the large scaled tonguefish Cynoglossus macrolepidotus (Bleeker). Proc. Natl. Inst. Sci. India 33:94-110. Lagler, K. F. 1956. Freshwater fishery biology, 2nd ed. W.C. Brown Co. Publ., Dubuque, lA, 421 p. Lin, M.-C. 1982. Age and growth of north Taiwan tongue sole Cynoglossus arel. Bull. Taiwan Fish. Res. Inst. 34:91-99. Lux, F. E. 1970. Note on growth of American plaice, Hippoglossoides platessoides iFabr) in ICNAF Subarea 5. Int. Comm. Northwest Atl. Fish. Spec. Publ. 7:5-7. 1973. Age and growth of the winter flounder Pseudopleuro- nectes americanus. on Georges Bank. Fish. Bull. 71:505-512. Lux, F. E., and F. E. Nichy. 1969. Growth of yellowtail flounder, Limanda ferruginea (Storer), on three New England fishing grounds. Int. Comm. Northwest Atl. Fish. Res. Bull. 6:5-25. Mansuetti, R., and R. Pauly. 1956. Age and growth of the northern hogchoker, Trinectes maculatus. in the Patuxent River. Maryland. Copeia 19.56:60-62. McEachron, L. W., G. C. Matlock, C. E. Bryan, P. Unger, R. J. Cody, and J. H. Martin. 1994. Winter mass mortality of animals in Texas bays. Northea.st Gulf Sci. 13(2): 121-138. Meng, T., and S. Ren. 1988. Age and growth of Cynoglossus semilaevis Giinther in the Bohai Sea. Mar Fish. Res. (Shandong) 9:173-183. [In Chinese, with English abstract.) Menon, A. G. K. 1977. A systematic monograph of the tongue soles of the genus Cynoglossus Hamilton- Buchanan (Pisces: Cyno- glossidae). Smithson. Contrib. Zool. 238:1-129. Miller, J. M., J. S. Burke, and G. R. Fitzhugh. 1991. Early life history patterns of Atlantic North Ameri- can flatfish: Likely (and unlikely) factors controlling recruitment. Neth. J. Sea. Res. 27:261-275. Moe, M. A., Jr., and G. T. Martin. 1965. Fishes taken in monthly trawl samples offshore of Pinellas County, Florida, with new additions to the fish fauna of the Tampa Bay area. Tulane Stud. Zool. 12:129-151. Moore, E. 1947. Studies on the marine resources of southern New England. 6. The sand flounder, Lophopsetta aquosa (Mitchill): a general study of the species with emphasis on age determination by means of scales and otoliths. Bull. Bingham Oceanogr. Coll. 11:1-79. Munroe, T. A. 1990. Eastern Atlantic tonguefishes (Symphurus: Cyno- glossidae: Pleuronectiformesl. with descriptions of two new species. Bull. Mar .Sci. 47:464-515. 1998. Systematics and ecology oi' tonguefishes of the genus Symphurus (Cynoglossidae: Pleuronectiformes) from the western Atlantic Ocean. Fish. Bull. 96:1-182. Murdy, E. O., R. S. Birdsong, and J. A. Musick. 1997. Fishesof Chesapeake Bay. Smithsonian Institution Press. Washington. DC. .324 p. Music, J. L., and J. M. Pafford. 1984. Population dynamics and life history aspects of ma- jor marine sportfishes in Georgia's coastal waters. Geor- gia Dcp. Nat, Res., Coa.st. Res. Div, Contr Scr 38:1-382. Olney, J. E., and G. C. Grant. 1976. Early planktonic larvae of the blackcheek tonguefish. Symphurus plagiusa (Pisces: Cynoglossidae), in the lower Chesapeake Bay Ches. Sci. 17:229-237. Parr, A. E. 1933. A geographic-ecological analysis of the seasonal changes in temperature conditions in shallow water along the Atlantic coast of the United States. Bull. Bingham Oceanogr. Lab. 4:1-90. Pitt, T. K. 1967. Age and growth of American plaice {Hippoglossoides platessoides) in the Newfoundland area of the northwest Atlantic. J. Fish. Res. Board Can. 24:1077-1099. Prather, E. E. 1967. The accuracy of the scale method in determining the ages of largemouth bass and bluegills. Proc. Ann. Conf Southeast. Assoc. Game Fish Commission. 20:483-486. Qasim, S. Z. 1973. Some implications of the problem of age and growth in marine fishes from the Indian waters. Indian J. Fish. 20(2):351-371. Rajaguru, A. 1992. Biology of two co-occurring tonguefishes, Cynoglossus arel and C. lida (Pleuronectiformes: Cynoglossidae), from Indian waters. Fish. Bull. 90:328-367. Ramanathan, N., P. Vijaya, V. Ramaiyan, and R. Natarajan. 1980. On the biology of the large scaled tongue sole Cynoglossus macrolepidotus (Bleeker). Indian .J. Fish. 24:83-89. Ramos, J. 1982. Estudio de la edad y crecimiento del lenguado, Solea solea (Linneo, 1758) (Pisces; Soleidae). Invest. Pesq. 46(l):15-28. Reichert, M. J. M., and H. W. van der Veer. 1991. Settlement, abundance, growth, and mortality of ju- venile flatfish in a subtropical tidal estuary (Georgia, U.S.A.). Neth. J. Sea. Res. 27:37.5-391. Ricker, W. E. 1975. Computation and interpretation of biological statis- tics offish populations. Department of the Environment, Fisheries and Marine Service, Fish. Res. Board Can. Bull. 191, Ottawa, Canada, 400 p, Robson, D. S., and D. G. Chapman. 1961. Catch curves and mortality rates. Trans. Am. Fish. Soc. 90:181-189. Roff, D. A. 1982. Reproductive strategies in flatfish: a first synthesis. Can. J. Fish. Aquat. Sci. 39:1686- 1698. Royce, W. F. 1972. Introduction to the fishery sciences. Academic Press, New York. NY, 351 p. Russell-Hunter, W. D. 1970. Aquatic productivity The Macmillan Co.. London, 306 p. Saila, S. B., C. W. Recksieck, and M. H. Prager. 1988. Basic fishery science programs. Elsevier, Amster- dam, 2.30 p. Santos, P. T. 1994. Growth and reproduction of the population of the four- spot megrim (Lepidorhombus boscii Risso) off the Portu- guese coast. Neth. J. Sea. Res. 32:379- 383. Seshappa, G. 1976. On the fishery and biologv- of the large tongue-sole, Cynoglossus dubius Day, at Calicut, Kerala. Indian J. Fish. 21:34.5-3.56. 1978. Report on a collection of tongue soles {Cynoglossus sp.) from Moplah Bay with a description of C. lida (Bleeker). Indian J. Fish. 23:160-173. Terwilliger and Munroe: Age, growth, longevity, and mortality of Symphurus plagiusa 361 1981. Observations on the size distribution and the occur- rence of growth rings in the scales of three species of Cynoglossus at CaUcut. Indian J. Fish. 25:188-196. Seshappa, G., and B. S. Bhimachar. 1954. Studies on the age and growth of the Malabar sole, Cynoglossus semifasciatus Day. Indian J. Fish. 1( 1 ):145- 162. 1955. Studies on the fishery and biology of the Malabar sole, Cynoglossus semifasciatus Day. Indian J. Fish. 2(1): 180-230. Seshappa, G., and B. K. Chakrapani. 1984. A scalimetric comparison of the malabar sole Cynoglossus semifasciatus Day from different centres of the west coast. Indian J. Fish. 31:82-89. Shealy, M. H., Jr., J. V. Miglarese, and E. B. Joseph. 1974. Bottom fishes of South Carolina estuaries — Relative abundance, seasonal distribution, and length frequency relationships. Tech. Rep. 6, South Carolina Marine Re- source Center, Charleston, SC, 189 p. Smith, R. W., and F. C. Daiber. 1977. Biology of the summer flounder, Paralichthys dentatus. in Delaware Bay Fish. Bull. 7,5:823-830. Spurr, A. R. 1969. A low-viscosity epoxy resin embedding medium for electron microscopy. J. Ultrastruct. Res. 26:31-43. Stickney, R. R. 1976. Food habits of Georgia estuarine fishes II. Symphurus plagiusa (Pleuronectiformes: Cynoglossidae). Trans. Am, Fish. Soc. 105:202-207. Terwilliger, M. R. 1996. Age, growth, and reproductive biology of the blackcheek tonguefish, Symphurus plagiusa (Cyno- glossidae: Pleuronectiformes), in Chesapeake Bay, Virginia. M.A. thesis. The College of William and Mary, Williamsburg, VA. 116 p. Toepfer, C. S., and J. W. Fleeger. 1995. Diet of juvenile fishes Citharichthys spilopterus, Symphurus plagiusa, and Gobionellus boleosoma. Bull. Mar Sci. 56( l):238-249. Topp, R. W., and F. H. Hoff Jr. 1972. Flatfishes (Pleuronectiformes). Fla. Dep. Nat. Resour, St. Petersburg. Mem. Hourglass Cruises 4:1-135. Vaughan, D. S., and M. L. Burton. 1994. Estimation of von Bertalanffy growth parameters in the presence of size- selective mortality: a simulated ex- ample with red grouper Trans. Am. Fish. Soc. 123:1-8. Victor, A. C. C. 1981. Length-weight relationship in the Malabar sole, Cynoglossus macrostomus Norman. Indian J. Fish. 25:259-262. Wada, K. 1970. Studies on the population biology of the flatfish, Limanda herzensteini Jordan et Snyder, in Nigata Region. Bull. Jpn. Sea Reg. Fish. Res. Lab. 22:31-43. Wenner, C. A., and G. R. Sedberry. 1989. Species composition, distribution, and relative abun- dance of fishes in the coastal habitat off the southeastern United States. NOAA Tech, Rep. NMFS-SSRF 79, 49 p. White, M. L., and M. E. Chittenden Jr. 1977. Age determination, reproduction, and population dy- namics of the Atlantic croaker, Micropogonias undulatus. Fish. Bull. 75:109-123. Zhu, X., and D. Ma. 1992. Studies on age, growth, and age class structures of the tonguefish. Cynoglossus abbreviatus (Day) in the Jiaozhou Bay waters. Mar Sci. (Beijing) 1:49-53. [In Chinese, with English abstract.] Zoutendyk, P. 1 974. The biology of the Agulhas sole, Austroglossus pecto- rails. Part-2: Age and growth. Trans. R. Soc. So. Afr 41(1): 33-41. 362 Abstract.— The lack of age and growth estimates and population parameters for the amberjack family prompted our study to describe otolith structure and determine size-at-age and growth rates of greater amberjack iSenola dumerili ) from the north-central Gulf of Mexico. Greater amberjack age and growth was described from a combination of external ridges and internal annuli in sectioned sagittal otoliths. A mark-re- capture study using tetracycline was consistent with a single annulus being formed each year in two- and three- year-old amberjack. Tag-recapture data from the Gulf of Mexico and South At- lantic Cooperative Gamefish Tagging Program also confirmed that a single annulus increment was formed each year. Ages for 597 specimens ranged from young-of-the-year to 15 years, and all fish over 9 years were female. Sag- ittal weight provided a reliable estima- tion of fork length with a relationship ofFL = 151 (1 _e i-oo4(SH'+i.6H). g ^gga. tive exponential equation in the form of the von Bertalanffy equation. The fit of the negative exponential in this application pro- vides evidence of the disassociation be- tween otolith growth and fish gi-owth as fish age. The von Bertalanffy growth curve equation, L, = 138.9 ( 1 -e-°-^"*° ™'), compared favorably with previous pub- lished values. Monthly changes in size and availability indicate a Gulf-wide mi- gration of this moderately long-lived pe- lagic predator. Age distribution and growth of greater amberjack, Seriola dumerili, from the north-central Gulf of Mexico* Bruce A. Thompson Marty Beasley Charles A. Wilson Coastal Fisheries Institute Center for Coastal, Energy, and Environmental Resources Louisiana State University Baton Rouge, Louisiana 70803-7503 E-mail address (for B A Thompson) coethoaunixl sncclsu edu Manuscript accepted 12 June 1998. Fish. Bull. 97:362-371 (1999). The greater amberjack, Seriola dumerili, is a pelagic reef species ranging along the American coasts from Nova Scotia to Brazil and oc- curring throughout the Atlantic, Pacific, and Indian oceans, as well as the Mediterranean Sea (Mather, 1958; Burch, 1979; Shipp, 1988). It is the largest member of the family Carangidae (Hoese and Moore, 1977). Commercial landings of amber- jack (all species) have increased markedly over the past twenty years. Landings in the Gulf of Mexico peaked in 1988 reaching 2.7 million pounds, with a dockside value over 1.6 million dollars, but have since declined to less than one million pounds in 1995-96. Recre- ational catch statistics are incom- plete but are suspected to have equalled or surpassed the commer- cial catch, raising concern over the status of the Atlantic stock in the late 1970s (Berry and Burch, 1979). Despite their popularity as a recre- ational and commercial quarry, little is known about the life histo- ries of amberjacks (Shipp, 1988). Of particular concern to stock- assessment scientists is the lack of age and growth estimates and popu- lation parameters for the ambeijack family (Burch, 1979; Humphreys, 1986). Burch (1979), using scales, aged greater amberjack from the western Atlantic to 10 years for fe- males and to eight years for males. Females were significantly longer than males beyond age four Hum- phreys (1986) reported von Berta- lanffy growth parameters for Ha- waiian greater amberjack but did not elaborate on how the fish were aged or how many fish were con- tained in the data set. The purpose of this study was to collect and describe greater amber- jack otoliths, assess their utility in age estimation, and determine the size-at-age and growth rates of greater amberjack from the north- central Gulf of Mexico. Materials and methods Sampling Greater amberjack («=840) were collected off the Louisiana coast from April 1989 to June 1992. Sources of samples included a com- mercial processor plant (7) = 18), charterboats (n=352), saltwater fishing tournaments (n =2 15), spear- fishing tournaments (7! = 159), recre- ational catches (n=48), and hook- ■ Contribution LSU-CFI-94-9 of the Coastal Fisheries Institute. CCEER. Louisiana State University. Baton Rouge. LA 70803- 7.503. Thompson et al.: Age distribution and growth of Senola dumenii 363 and-line catches from an ofFshore gas production plat- form (;i=48). In general sampling was not system- atic; we examined all fish available. Morphometric measurements and otoliths were collected. Fish were measured by using fork and to- tal lengths (FL and TL> in mm and weighed by using total and gutted weights in g. Both sagittae were removed and stored in 95*^ ethanol for later exami- nation. Lapilli, asterisci, and dorsal spines were re- moved from several fish to evaluate their usefulness for age estimation. Otoliths were cleaned of organic tissue by rinsing in a 50% hypochlorite solution iClorox), air dried, weighed to the nearest 0.1 mg with a Sartorius model 1801 microbalance, and stored dry (Wilson et al.,1991). Age estimates were made by using techniques simi- lar to those of Wilson et al. (1991) for billfish. Whole otoliths were sputter coated with a mixture of gold and palladium and viewed with reflected light un- der a dissecting microscope for description of exter- nal morphology. These samples were also viewed with a Cambridge Stereoscan 150 scanning electron mi- croscope (SEM) to obtain detailed photographs at niagnifications ranging from 20 to 650x. Sagittae were embedded in epoxy resin (Spurr, Kmbed 812, or Araldite GY 502) and sectioned with a Buehler Isomet low-speed saw to yield a thin (about 1-mm) transverse section containing the core I Beckman, 1989). Transverse sections were ground with various grades (300-2000 grit) of wet and dry .'sandpaper until the core was at the surface, then polished with 0.3 |.im alumina polish. Thin sections were mounted on glass slides with a clear thermo- plastic cement (Crystalbond 509) and viewed under transmitted light with an Olympus BH-2 compound microscope at 50-250x magnification. Age estimates from sectioned sagittae were made by combining counts of translucent and opaque zones and other growth features viewed under transmitted light and counts of associated ridges on the ventral and medioventral portions of the rostrum as determined in billfish species (Wilson and Dean, 1983; Prince et al., 1986; Wilson et al., 1991). Validation and verification of our age estimation technique were attempted by using several tech- niques. Validation of annulus formation using mar- ginal increment analysis proved futile owing to our inability to determine the condition of the otolith edge. Therefore we pursued additional techniques that would corroborate our age estimates. A mark- recapture study was carried out from a gas produc- tion platform (Mobil USA, West Cameron block 352) located in the Gulf of Mexico 50 miles south of Cameron, Louisiana. Greater amberjackt « =48) were captured by hook and line, measured as described above, tagged at the base of the dorsal fin with a Hall- print dart tag, injected with oxytetracycline hydro- chloride (Agrimycin-100) at a rate of 20-40 mg/kg fish weight, and released. Tagged greater amberjack (6) were recaptured with hook and line or by spear fish- ermen at the release site and sampled as described above. Sectioned sagittae were viewed under ultra- violet light (405-435 nm wavelength) at 50-250x magnification for detection of the tetracycline mark. Attempts made to compare age estimates from sag- ittal otoliths with dorsal and anal spines and verte- brae were not successful. To determine reproducibil- ity of age estimates, sagittae were aged indepen- dently by two readers. Reproducibility of age esti- mates was compared by using the coefficient of varia- tion (CV) and index of precision (D) (Chang, 1982). Tag-recapture data for the Gulf of Mexico and South Atlantic (7!=711) were obtained from the Co- operative Gamefish Tagging Program (CGTP). Fish that had been at large for at least 365 days after tagging and that showed positive growth were used. Twenty-five specimens met these criteria. Following the method of Labelle et al. (1993), we used the von Bertalanffy growth curve parameters to predict changes in length from the tag recapture data to verify our age estimates. Like Labelle et al. (1993) we employed Fabens' ( 1965) length increment model to predict length at recapture: lr,=l,+{L ■/,)(l-e-*^'), where l^ At Ir length at release of individual i; time of liberty of individual /; estimated length-at-recapture of indi- vidual ;; and L^ and k = von Bertalanffy parameters estimated from otolith ages. Predicted recapture lengths were then compared with observed recapture lengths. As Labelle et al. (1993) pointed out, this procedure is not statistically rigor- ous but is useful for comparative purposes. An analysis of variance (ANOVA) was used to test for sex-related differences in mean fork length, mean gutted weight, mean sagittal weight, and mean age among sources and within age class. Analysis of co- variance (ANCOVA) was used to test for differences in the relations (fork length [L] versus total weight [Wt], fork length versus sagittal weight [SW], age versus sagittal weight) between sexes (Cerrato, 1990; Kimura, 1980). Statistical inferences were made with a significance level of a=0.05. The relationship be- tween fork length and otolith weight was modeled with a negative exponential (von Bertalanffy) be- cause it provided the best fit (highest /•-). 364 Fishery Bulletin 97(2), 1999 The von Bertalanffy model L, = LJ 1 - e"*" " '»') was also used to describe the relation between length and age. Relative ages were assigned to fish by using a birth date of 1 April based on trends in gonadoso- matic indices and the regression of sagittal weight on month of capture for young-of-the-year fish.^ Von Bertalanffy growth parameter estimates for males and females were compared by using a likelihood ratio test at a = 0.05 (r statistic) (Cen-ato, 1990; Kimura, 1980). Results Length and weight Greater amberjack ranged from 167 to 1441 mm (FL); females ranged from 374 to 1441 mm ( .r =879 mm), and males ranged from 387 to 1203 mm ( .v =854 mm). Females represented 72% of the fish over 1 m in fork length. There was no significant difference in mean fork length between males and females. Females ranged from 0.82 to 42.5 kg (.r=11.33) and males ranged from 0.85 to 28.8 kg ( .r =9.90). Mean gutted weight (GW) of females was significantly gi'eater than that of males (ANOVAP>F= 0.03) and females repre- sented 789c of the fish over 20 kg gutted weight. Re- gi'essions for TL versus FL and TW versus GW were Table 1 Monthly mean fork length ( mm l and se.x ratio of charterboat- caught greatei amberjack from the northern Gulf of Mexico 1 ranked by decreasing mean fork length. Mean fork Sex ratio Month length (mm) n female:male May 771.1 70 2.0 Jul 702.7 11 1.6 Jun 651.0 21 2.7 Sep 613.2 14 0.9 Mar 566.9 75 2.3 Apr 560.4 43 3.5 Nov 482.6 38 2.0 Oct 420.8 59 2.0 Feb 408.4 9 5.0 Dec 383.0 9 0.0 Jan 351.5 2 0.0 Females: Males: Sex unknown: W= 3.25 X 10-5 ^2 87 (,.2=0.98, n=324) W =1.75x10-^1^^^ W=7.62xlO-5L-8' (n=0.98,n = 184i (r2=0.98, «=354) Otolith structure TL = 1.14 i^L + 13.05 TW = 1.09 GW+ 119.40. (r'-=0.99) (r-=0.99) Monthly mean fork length and sex ratio from charterboat catches were compared because this was the only year-round source of greater amberjack (Table 1). Charterboats fishing Gulf of Mexico wa- ters off the Louisiana coast caught larger greater amberjack during the summer months (from May to September) and smaller greater amberjack during winter months (from November to February). Sex ratios ranged from 0.4:1 to 4.5:1 (x=2.49) (Table 1), and there was a relatively greater abundance of fe- males each month except September. The relation between fork length (cm) and gutted weight (kg) was best described by a power function. The slopes of the regression lines for males and fe- males were significantly different. The equations explaining these relations were All samples: W= 4.2 x 10-^ L-'^'' (7-2=0. 99. «=862) ' Thompson, B. A.. C. A. Wil.son, .) H KcndiT, M Beasley, and C. Cauthron. 1992. Age, growth, and reproductive biology of greater amberjack and cobia from Louisiana waters. Final re- port to U.S. Department of Commerce, Marine Fisheries hiitia- tive (MARFIN) Program, NMFS, St. Petersburg, FL, NA90AA-H- MF722, 77 p. Greater amberjack sagittae are small, thin, fragile, and elongate in the anterior direction and bluntly crenelate at the posterior end. The medial surface is convex and has a deep, prominent sulcus. The ante- rior portion of the sagitta is cui-ved laterally and the posterior end is relatively flat. The rostrum is longer than the antirostrum, but the difference increases with fish size. Prominent grooves and ridges are present on the lateral side of the sagittae and are nearly absent on the medial side (Fig. 1). The asteriscus and lapillus were much smaller and more fragile than the sagittae and have no annuluslike external or internal growth features. The small size and fragility of these smaller otoliths, as well as the time and care required to remove them, precluded additional sampling. Greater amberjack sagittae weights ranged from 1.4 to 68.6 mg. The mean weight of male sagittae (23.2 mg) was not significantly different from that of females (25.0 mg). The relationship between FL and sagittal weight (SW) was best described with a nega- tive exponential (following the von Bertalanffy equa- tion). Because there was no difference in this rela- tion between males and females {P>x^>0.Q5 ) the data were combined to produce the equation (Fig. 2) FL s\v 151 (1-e -0 04iS\V +1.61 1 V (r2=0.96) Thompson et a\ Age distribution and growth of Senola dumen/i 365 Figure 1 Scanning electron micrograph of a sagitta (lateral view) (A), lapillus (B), and asteriscus (C) from a 42.7-cm greater amberjack from the north-central Gulf of Mexico. where FL^^ = fork length (cm); and SW = sagittal weight (mg). Although internal annuluslike features were visible in whole otoliths, they were not sufficiently translucent to permit con- sistent enumeration. Readability was not improved by immersing sagittae in either glycerin or clove oil for up to one year. Surface grooves and ridges were promi- nent on the lateral side of sagittae from young fish but became compressed and indistinguishable in older specimens ow- ing to an over-burden of calcium carbon- ate; a similar observation was reported for marlin (Wilson et al., 1991). External ridges were closely associated with inter- nal microstructure that we interpreted as annuli. Although a first and second annu- lus was visible in whole sagittae, we con- cluded that annuli could not be enumer- ated from whole otoliths for most fish, par- ticularly larger specimens. Microstructural growth features were readily vis- ible in most sectioned sagittae; however, the only consistent annuluslike features were regions where 160 1 140 120 .^ ' ••'■:-':'^''''-" ■ • • Fork length (cm) § § 8 J^ 40 ■Jr 20 f n 0 10 20 30 40 50 60 70 80 Sagittal weight (mg) Figure 2 Sagittal weight ( mg) versus fork length I cm 1 for north-central Gulf of Mexico greater amberjack. .v represents predicted values from given equation. growth bands converged on the lateral side of the otolith section. These "convergence zones" were in- ternal to the lateral surface of the antirostrum, where numerous growth features (increments) converged 366 Fishery Bulletin 97(2), 1999 on a common point (Fig. 3A); these features were usually associated with ridges on the lateral side of the sagitta or associated with alternating opaque and translucent zones, or with both ridges and zones (Fig. 3B). Similar structures were described and used to age istiophorids by Prince et al. ( 1986) and Wilson et al. (1991). Six of 48 oxytetracycline-injected fish were recov- ered from the tag-recapture experiment. Oxytetra- cycline marks were observed in sectioned sagittae from all six fish (Fig. 4). The location of the oxytetra- cycline mark in relation to presumed annuli sup- ported the use of the gi-owth features described above. Figure 3 Photomicrographs of sectioned sagittae from a 72-cm-FL (A) and 42-cm-FL (B) north-central Gulf of Mexico greater amberjack. Photomicrograph A shows the convergence zones (C) along the dorsolateral edge of the rostrum. Photomicro- graph B shows the opaque and translucent zones in the rostrum; opaque zones were interpreted as annuli (Al. The specimen in Figure 4 was captured in Septem- ber and recaptured in March and had an annulus consisting of a ridge and convergence zone between the tetracycline mark and the margin. The combined information obtained from six oxytetracycline- marked otoliths confirmed that one annulus is formed between November and March in the sagittae of two- and three-year-old greater amberjack. Age structure Five hundred and ninety seven greater amberjack sagittae were sectioned and aged by two readers. Forty-five specimens were consid- ered unreadable owing to poor sample preparation. Pairwise com- parisons between two readers in- dicated that 9.6% of the counts dif- fered by 1 year, 1.6% differed by 2 years, and 0.4% differed by more than 2 years. Readers combined had a coefficient of variation (CV) of 0.15 and an index of precision (D)ofO.ll. Age estimates of greater amber- jack ranged from young-of-the- year to 15 years. Mean age of males (2.9 yiO was not significantly different than that of females (3.2 yrXANOVA P>F=0.0641; however, all fish over 9 years of age were female. Most of the greater amber- jack sampled were from one to three years old (Fig. 5). A linear function best described the relation between estimated age and sagittal weight. The regres- sion parameters were not signifi- cantly different (ANCOVA) be- tween males and females; there- fore data were pooled and pro- duced the relation Age = 0.15(SW) - 0.44, (r-^=0.75) where SW = sagittal weight (mg); and Age = amberjack age in years. Growth was modeled by using the von Bertalanffy growth model (Fig. 6) to facilitate comparisons with other values reported in the literature. Our age data from Thompson et al.; Age distribution and growth of Senola dumerili 367 annuli Figure 4 Photomicrograph of a sectioned sagitta IVom a 53-cm tagged and recaptured north- central Gulf of Mexico greater amberjack obser\'ed under ultraviolet light. The fish was at large for 160 days. Tetracycline mark = m; annuli are indicated by the arrows. greater amberjack were fitted to the von Bertalanffy growth model (Fig. 7). Because there was no differ- ence between male and female models (P>x^=0.496), data were combined and resulted in L, - 138. 9( 1 - e-° 25't ^ 0.791) (^2=0.96, n - 552) Tag-recapture data obtained from the CGTP yielded 25 data points useful for comparison of growth rates, and an additional six data points from our mark-re- capture experiment. Predicted lengths fell along the trajectory of a line where predicted lengths equalled observed lengths (Fig. 7). The differences between observed and predicted values were both positive and negative. There was a tendency to under-predict fish size above 80 cm recapture length. The growth of tagged and recaptured fish provided additional sup- port of our age estimation technique. Discussion Our observations of sex-related differences in maxi- mum size of greater amberjack are consistent with the findings of previous investigators. Burch (1979) reported a maximum size of 145.0 cm for males and of 155.5 cm for females and noted that the mean fork 200 175 150 >. 125 o § 100 £ 75 u. 50 25 0 I I Females Q Males n Sex unknown <1 4 6 8 Age (yr) 10 12 14 Figure 5 Estimated age-frequency histogram, identified by sex, of northern Gulf of Mexico greater amberjack collected from April 1989 to June 1992. length of females was greater than that of males of the same age. Humphreys (1986) reported a maxi- mum size of 106.0 cm for males and of 149.4 cm for females. We found that the maximum size of males (132.7 cm) was less than that of females (144.1 cm) and that females represented 72% of the fish over 368 Fishery Bulletin 97(2), 1999 100 cm. Our data support the hypotheses that the maximum size in greater amberjack is sex related. Similar results were reported for other pelagic spe- cies (Atlantic and Pacific blue marlin, and swordfish) suggesting that females either grow faster, live longer, or both (Wilson, 1984; Wilson et al., 1991). Charterboat-caught greater amberjack were col- lected year-found from the Louisiana coast, provid- ing us with the opportunities to examine seasonal changes in the population. Size-frequency analysis of these data showed that fish were largest from May to September and smallest from November to Feb- Estimated age (yr) Figure 6 Plot of the estimated age and length of north-central Gulf of Mexico greater amberjack and the resultant von Bertalanffy relationship. ruary, suggesting a seasonal shift in the size fre- quency of individuals in the population off Louisi- ana. Although this seasonal size difference could be due to sampling bias, we believe that greater amber- jack have an affinity for warmer waters and that the cooler waters associated with winter initiate an emi- gration from Louisiana waters. Burch ( 1979 ) reported greater ambeijack migrating southward from Decem- ber to May and northward from June to December in Florida. Baxter (1960) reported a northward mi- gi'ation of California yellowtail (as Seriola dorsalis) off California waters during early spring. The size at which greater amberjack begin to mi- grate is not known (Burch, 1979). Females tended to be more abundant than males throughout our study, although the female-to-male ratio varied with time of year and source. This is contrary to Burch ( 1979) who reported female-to-male ratios from 0.6:1 to 2:1 with males more numerous in all months except July, Au- gust, and September Humphreys (1986) reported a female-to-male ratio of 0.9:1. In our study, no trend in sex ratio was de- tected for charterboat-caught greater am- berjack, and the overall ratio was 2.5:1. It is possible that males are more abundant off the west coast of Florida and females are more abundant off Louisiana. Thomp- son et al.^ found that male cobia iRachy- centron canadum) dominated in the west- central Gulf of Mexico, whereas females dominated in the east-central Gulf of Mexico. 20 40 60 80 100 Predicted values (cm) 120 140 160 Figure 7 Predicted versus ob.served recapture lengths of tagged greater amber- jack. Predicted values were calculated by using our von Bertalanffy growth parameters. The solid line covers points where the predicted and observed values are equal. Otoliths Greater amberjack otoliths are small, frag- ile, and complex in shape. In a compari- son of external and internal features with those of other pelagic species, sagittae were found to be similar in shape to those of red steenbras (Petriis rupestris) (Smale and Punt, 1991), tunas (C.A. Wilson, unpubl. data), billfishes (Wilson et al., 1991), and several other carangids (Alectis, Caranx, Chloroscommbrus, Elegatis, and Uraspis, senior author, unpubl. data) We tried several models (power function, logistic equation, Gompertz and Richards models) to describe the relationship be- tween fish size (fork length) and otolith size (sagittal weight). A negative exponen- tial in the form of the von Bertalanffy equa- tion provided the best fit (r^=0.96) and in- dicated that sagittal weight continues to Thompson et al : Age distribution and growth of Seriola dumerili 369 increase even after increases in fish length or weight slows or ceases (Pawson, 1990). This finding provides further evidence that otolith growth is more closely associated with time than with fish size as the fish becomes older. Secor and Dean ( 1989) suggested that otolith growth was a function of both somatic growth and time in young fish. We propose that as fish grow and approach an asymptotic size, otolith growth con- tinues at a basal level and does not reach an asymp- tote. This is further evidence of the disassociation of otolith growth and fish growth in older fish. Contin- ued sagittal growth throughout the life of the fish is the property that makes sagittae a useful ageing tool (Gamboa, 1991; Casselman, 1990). Opaque zones (annuli) observed in transverse sec- tions of greater amberjack sagittae were not as con- sistent as those reported in other species such as the sciaenids (Beckman, 1989; Casselman, 1990). Wil- son et al. (1991) found that establishment of annual patterns in transverse sections of sagittae from pe- lagic fishes was best accomplished through interpre- tation of a combination of growth features on the surface of and in sectioned sagittae. Transverse sec- tions of greater amberjack sagittae contained growth features described by Smale and Punt ( 1991) for red steenbras and by Wilson (1984) and Prince et al. ( 1986) for billfishes. We used similar criteria for dis- crimination of annual patterns in transverse sections of greater amberjack sagittae. Our criteria included external ridges on the lateral side of the sagittae, ridges within the sulcus acousticus, internal opaque and translucent zones, and the association of these features with a common area where many growth bands converged (Fig. 3A). All these growth features were coincident with annuli that were validated by using oxytetracychne-injected fish (Fig. 4). Smale and Punt (1991) found similar patterns in transverse sections of sagittae from red steenbras (Petrus nipestris). Although they reported fish as old as 25 years, transverse sections of sagittae fi-om red steenbras looked similar to those of greater ambeijack. Disagreement over age estimates based on the above criteria was usually over an assignment of the first annulus. In some specimens a prominent opaque zone occurred adjacent to the core; we concluded it was laid down during the first winter and therefore did not consider it an annulus. Baxter ( 1960) reported that the first annulus was formed on the scales of yellowtail (as Seriola dorsalis) during their second winter, which was 18 months after hatching. We con- sidered the first annulus in amberjack to be that which was laid down in the second year, when fish were actually 15 to 21 months old. When corrected for the discrepancy of the first annulus, the coeffi- cient of variation and index of precision for the two readers were 0.15 and 0.11, indicating reasonable consistency in annulus interpretation. Similar val- ues were reported by Wilson et al. (1991) for Atlantic blue marlin. Apparently the sex-related size difference in am- berjack is due to age-related differential mortality. Most of the greater amberjack collected were under 5 years old, although we found fish as old as 15 years. Males ranged from 1 to 9 and females ranged from 1 to 15 years. Females represented 76% of the samples above age 7. Burch ( 1979) found female greater am- berjack to age 10 and males to age 8 for fish taken by southern Florida charterboat fishermen. The only other pelagic predator that demonstrates age-related sexual dimorphism, where females outlive males, is swordfish (Wilson and Dean, 1983). Greater amber- jack also share a life history similar to both great barracuda and cobia in that they are all moderately long-lived pelagic reef species, and greater amber- jack could be expected to exhibit similar patterns of longevity for males and females. However, deSylva (1963) found no age difference between sexes for great barracuda (Sphyraena bar-racuda). Richards (1967) reported no sex-related age differences for cobia (Rachycentron canadum ) from mid-Atlantic waters, and Thompson et al.^ reported similar findings for cobia from the Gulf of Mexico. We conclude that greater amberjack males die at a younger age than do females. More research is needed to determine whether these differences are real or are the result of sampling biases used in our study. The relation between sagittal weight and age dem- onstrated that otolith weight can be used to estimate relative age. This finding is consistent with that for other pelagic species (Wilson, 1984; Wilson et al., 1991 ). Sagittal weight can provide a useful manage- ment tool because the random sampling and weigh- ing of greater amberjack sagittae can be used to es- timate age without the additional time and cost re- quired for sectioning. We used a negative exponen- tial to model the relationship between otolith size and fish size. The fit of this model indicates that otolith size continues to increase over time. Growth The von Bertalanffy growth parameter estimates obtained in our study were similar to both Burch's (1979) and Humphreys' (1986). Like Humphreys (1986), we did not find differences by sex. Burch ( 1979) reported L„ of 146.3 cm, 159.7 cm, and 164.3 cm for males, females, and sexes combined for charterboat- caught greater amberjack in Florida waters. Humphreys (1986) reported L^ of 149.3 for Hawaiian greater amberjack (sexes combined). We 370 Fishery Bulletin 97(2), 1999 obtained an Ljai 138.8 cm, slightly smaller than that obtained by other authors; differences may be due to duration of the growing season, because greater amberjack studied by other authors came from warmer climates where growth may have been more rapid. Our comparison of observed recapture lengths with predicted recapture lengths (Fig. 7 ) further veri- fied our age-estimation technique. Of interest was our observation that this method tended to under- estimate fish size at sizes larger than 80 cm. Because most of the tagged fish were collected off south Florida, it is possible that they grow slightly faster than those off Louisiana. Conclusion Our analysis provides evidence that greater amber- jack from the north-central Gulf of Mexico are moder- ately long-lived, living up to 15 years. They demonstrate a sexual dimorphism, where females grow larger than males, but this appears to be age related because males die younger. Our use of sagittal annuli is vali- dated for only two- and three-year-old fish and future studies will be needed to complete age validation. Acknowledgments Funding for this research was provided by the U.S. Department of Commerce Marine Fisheries Initia- tive (MARFIN) Program and Louisiana State LTni- versity Coastal Fisheries Institute. This study was originally submitted as an M.S. thesis to the Depart- ment of Oceanography and Coastal Sciences, Louisi- ana State University, by Marty Beasley. We thank David and Louise Stanley for their field and laboratory assistance. We also thank the staff of Mobil Oil Co. for their support in the work done at the West Cameron 352 platform, Charlie Hardison for access to his charterboat catches, and the many staff and fishermen of the Louisiana saltwater fish- ing rodeos for assistance in sampling greater amber- jack entered in the rodeos, and Jay Geahgan for his statistical assistance. Literature cited Baxter, J. L. 1960. A study of the yullowtail, Senula dorsalis (Gill). Dep. Fish Game Bull. 110. 96 p. Beckman, D. W. 1989. Age and growth of red drum. Sciaenops ocellatus. and black drum, Pogonias cromis. in the northern Gulf of Mexico. Ph.D. diss., Louisiana State Univ., Baton Rouge, LA, 161 p. Berry, F. H.. and R. K. Burch. 1979. Aspectsof the amberjack fisheries. /nJ.B. Higman (ed.) Proceedings of the thirty-first annual Gulf and Car- ibbean Fish, Inst,, Miami, FL, p. 174-179. Burch, R. K. 1979. The greater amberjack, Seriola dumenli: its biology and fishery off southeastern Florida. M.S. thesis, Univ. Miami, Miami, FL. 112 p. Casselman, J. M. 1990. Growth and relative size of calcified structures of fish. Trans. Am. Fish. Soc. 119:673-688. Cerrato, R. M. 1990. Interpretable statistical tests for growth comparisons using parameters in the von Bertalanffy equation. Can. J. Fish, Aquat, Sci, 47:1416-1426. Chang, W. Y. 1982. A statistical method for evaluating the reproducibil- ity of age determination. Can, J, Fish, Aquat. Sci. 39:1208-1210. deSylva, D. P. 1 963. Systematics and life history of the great barracuda, Sphy- raena barracuda. Stud, Trop, Oceanogr, (Miami! 1:1-179. Fabens, A. J. 1965. Properties and fitting of the von Bertalanffy growth curve. Growth 29:26.5-289, Gamboa, D. A. 1991. Otolith size versus weight and body-length relation- ships for eleven fish species of Baja California, Mexico, Fish, Bull, 89:701-706. Hoese, H. D., and R. H. Moore. 1977. Fishes of the Gulf of Mexico: Texas, Louisiana, and adjacent waters, Texas A&M Univ, Press, College Station, TX, 327 p, Humphreys, R. L. 1986. Carangidae — greater amberjack. In R, N, Uchida and .J, H, Uchiyama leds,). Fishery atlas of the northwest- ern Hawaiian Islands, p, 100-101, US, Dep, Commer, NOAA Technical Report, NMFS 38, Kimura, D. K. 1980. Likelihood methods for the von Bertalanffy growth curve. Fish. Bull, 77:76.5-776 Labelle, M., J. Hampton, K. Bailey, T. Murray, D. Fournier, and J. Sibert. 1993. Determination of age and growth of South Pacific albacore '■Thunnus alalunga) using three methodologies. Fish. Bull. 91:649-663. Mather, F. J. 1958. A preliminary review of the amberjacks, genus Seriola, of the western Atlantic. In Proceedings of the third international game and fish conference, p, 1-13, Pawson, M. G. 1990. Using otolith weight to age fish, J, Fish Biol, 36: 521-.531. Prince, E., D. W. Lee, C. A. Wilson, and J. M. Dean. 1986. Longevity and age validation of a tag- recaptured Atlan- tic sailfish. Istiophorus platyptcriis. Fish. Bull, 84:493-502, Richards, C. E. 1967. Age, growth and fecundity of the cobia. Rachycentron canadum. from Chesapeake Bay and adjacent mid-Atlan- tic waters. Trans. Am, Fish, Soc, 96:343-350, Sccor, D. H., and J. M. Dean. 1989. Somatic growth effects on the otolith-fish size rela- tionship in young pond-reared striped bass, Morone sa.ta- tili.i. Can. ,1 Fi.sh. Aquat. Sci. 46:113-121. Shipp, R. L. 1988. Dr Bob shipp's guide to fishes of the Gulf of Mexico. Century Printing, Mobile, AL, 256 p. Thompson et dl.: Age distribution and growth of Seriola dumerili 371 Smale, M. J., and A. E. Punt. 1991. Age and growth of the red steenbras Petrus rupestris (Pisces: Sparidae) on the south-east coast of South Africa. S. Afr. J. Mar. Sci. 10:131-139. Wilson, C. A. 1984. Age and growth aspects of the life history of billfishes. Ph.D. diss., Univ. South Carohna, Columbia, SC. 180 p. Wilson, C. A., J. M. Dean. 1983. The potential use of sagittae for estimating age of Atlantic swordCish, Xiphias gladius. In E. D. Prince and L. M. Pulos (eds.). Proceedings of the international work- shop on age determination of oceanic pelagic fishes: tunas, billfishes. and sharks, p. 1.51-156. U.S. Dep. Commer., NOAA Tech. Rep. NMFS 8. Wilson, C. A., J. M. Dean, E. D. Prince, and D. W. Lee. 1991. An examination of the sexual dimorphism in Atlan- tic and Pacific blue marlin using body weight, sagittae weight, and age estimates. J. Exp. Mar. Biol. Ecol. 151:209-225. 372 Mercury levels In four species of sharks from the Atlantic coast of Florida Douglas H. Adams Florida Marine Research Institute Florida Department of Environmental Protection 1220 Prospect Ave , Suite 285 Melbourne, Florida 32901 E mail address adams_dti gepic7dep state fl us Robert H. McMichael Jr. Flonda Marine Research Institute Florida Department of Environmental Protection 100 Eighth Ave SE St. Petersburg, Florida 33701 Florida's commercial and recre- ational shark landings represent a significant portion of the total U.S. Atlantic shark landings (NMFS, 1993). Shark landings have in- creased significantly during the past decade because human con- sumption of shark meat has become increasingly acceptable and be- cause, in Asian markets, the de- mand for shark fins is very high — as are the prices paid for them (NMFS, 1993; Brown, in press). The east-central coast of Florida is an important area for commercial and recreational shark fishing, and a wide array of shark species, includ- ing those examined in this study, are landed in this region (Trent et al., 1997; FDEP'). Mercury, a toxic metallic ele- ment, has been shown to bioac- cumulate in fish tissue, and there- fore, fish can represent a major di- etary source of mercury to humans (Phillips and Buhler, 1978; Turner et al., 1980; Lyle, 1986). Methyl- mercury is the most toxic form of mercury for humans to consume (Meaburn, 1978; NMFS, 1993) and essentially all mercury found in fish muscle tissue (>95'}f ) is in the monomethyl form (CH,3Hg)(Grieb et al., 1990; Bloom, 1992). There- fore, the measurement of total mer- cury provides an approximation of methylmercury and has been rec- ommended as the standard for regulatory monitoring (Bloom, 1992). Elevated mercury concentra- tions in fish have been a growing concern among resource manage- ment agencies. Apex predators, par- ticularly long-lived species such as billfishes (Forstner and Wittman, 1981; Barber and Whahng, 1983; Kai et al., 1987), tunas (Miller et al., 1972), mackerels (Meaburn, 1978). and sharks (Forrester et al., 1972; Walker. 1976; Lyle, 1986; Vas, 1991; Hueter et al., 1995, and oth- ers ) have been reported to accumu- late relatively high levels of mercury. In May 1991, the Florida Depart- ment of Health and Rehabilitative Services (FHRS) released a health advisory urging limited consump- tion of all shark species from Florida waters." Owing to mercury concentrations in excess of U.S. Food and Drug Administration and State of Florida standards, FHRS recommended "adults should eat shark no more than once a week; children and women of childbear- ing age should eat shark no more than once a month." State of Florida guidelines recommended that fish containing less than 0.5 ppm of to- tal mercury should represent no dietary risk, fish containing 0.5 to 1.5 ppm of total mercury should be consumed in limited amounts, and fish containing greater than 1.5 ppm of total mercury should not be consumed. The 1991 health advi- sory regarding sharks in Florida waters was derived from a limited number of samples taken from re- tail sources and from studies that lacked important information re- garding species, capture location, sex, and size of sharks examined. Increased landings of sharks in Florida for human consumption (Brown, in press; FDEPM has prompted the need for more de- tailed information regarding mer- cury levels in Florida shark species. Consequently, we report here analyses of total mercury levels in the muscle tissue of three carcha- rhinids (bull shark, Carcharhinus leiicas; blacktip shark, C. limbatus; and Atlantic sharpnose shark, Rhizo- prionodon terraenovae) and one sphymid (bonnethead shark, Sphyr- na tibiiro ) from the east-central coast of Florida. Materials and methods Sample collection and mercury analysis Sharks were collected during the Florida Department of Environ- mental Protection, Florida Marine Research Institute's Fisheries-In- dependent Monitoring Program in the Indian River Lagoon system and adjacent coastal waters or from commercial gillnet or longline fish- eries operating in the nearshore ' FDEP I Florida Department of Environ- mental Protection) Marine Fi.sheries In- formation System. Fi.sheries Assessment Section. 100 Eiglith Ave. SE, St. Peters- burg, FL 33701. - FHRS (Florida Department of Health and Rehabilitative Services). 1991. Health Advisory for Marine Fish. 1317 Winewood Blvd.. Tallahassee. FL 32399, 3 p. Manusci-ipt accepted 20 May 1998. Fish. Bull. 97:372-379 1 19991, NOTE Adams and McMichael: Mercury level in four species of sharks from the Atlantic coast 373 and offshore waters of east-central Florida. All bull sharks were collected from estuarine nursery habi- tats within the Indian River Lagoon system from Oak Hill, Florida (approximately 28°52'N), south to the Ft. Pierce Inlet area (approximately 27"28'N). All blacktip, sharpnose, and bonnethead sharks were collected in nearshore and offshore waters from northern Cape Canaveral, Florida (approximately 28°55'N), south to the Sebastian InletAVabasso, Florida, area ( approximately 27°45'N ). Samples were collected from 1992 to 1995. Sharks were placed directly on ice and returned to the laboratory, where species, precaudal length (PCD, and sex were recorded. Stage of maturity was determined by examination of internal and external reproductive organs, as well as by comparison of shark size with estimates of size at birth or matu- rity from previous studies (Parsons, 1983, 1993; Branstetter and Stiles, 1987; Castro, 1996). Individu- als were classified as neonates on the basis of un- healed or healing umbilical scars (Castro, 1993). To allow for comparisons with other studies, total length (TL), fork length (FL), and other morphometries were also recorded according to Compagno (1984). With the exception of embryos, all sharks sampled were considered to be within the size range landed in Florida recreational or commercial fisheries. Axial muscle tissue samples were removed from the left dorsal area anterior to the origin of the first dorsal fin. White muscle tissue taken from this region is representative of the portion of shark used for hu- man consumption. Tissue samples were immediately placed in sterile polypropylene vials, sealed, and fro- zen at -20°C until analyzed. Before analysis, tissue samples were digested by using standard procedures (EPA, 1991; Frick'^) to convert all mercury in the sample to Hg^-. The mercury in each digested sample was reduced to atomic mercury by reaction with ex- cess stannous chloride. This atomic mercury was purged from solution in a gas-liquid separator and swept into an atomic absorption spectrometer for detection and quantification following standardized procedures (EPA, 1991; Booeshahgi et al.^) at the Florida Department of Environmental Protection's Division of Technical Services by using cold vapor atomic absorption spectrometry. Quality control mea- ■^ Frick.T. 1996. Digestion of fish tissue samples for total mer- cury analysis. Florida Department of Environmental Protec- tion. Division of Technical Services, 2600 Blair Stone Road, Tallahassee. FL 32399. Report MT-015-1. ^ Booeshahgi, F, M. Witt, and K. Cano. 199.5. Analysis of to- tal mercury in tissue by cold vapor atomic absorption. Florida Department of Environmental Protection. Division of Techni- cal Services, 2600 Blair Stone Road, Tallahassee, FL 32399. Report MT-010-1. sures included analysis of laboratory blanks, dupli- cate or triplicate tissue samples, and standard fish tissue reference material (DORM-1, obtained from the National Research Cotmcil of Canada) for each group of ten shark samples analyzed (EPA, 1991; Frick'^; Booeshahgi et al,"*). All mercury levels are reported as parts per million (ppm) wet weight. Data analyses Data regarding size and total mercury for each spe- cies were tested for normality by using the Kolmo- gorov-Smirnov test with Lilliefors' correction (Fox et al., 1994) and for homoscedasticity by using the F,,,^,- test (Sokal and Rohlf, 1981) or by computing the Spearman rank correlation between the absolute values of the residuals and the observed value of the dependent variable (Fox et al., 1994). Linear regres- sions were used to describe relationships between shark size and total mercury concentration. Mercury data were log transformed to meet homoscedascity requirements. We used a ?-test or Mann-Whitney rank sum test, as appropriate, to test for significant differences in total mercury levels and sizes between males and females for each species. Results Bull shark, Carcharhinus leucas Eighty-three percent of 53 neonate and juvenile bull sharks (552-1075 mm PCD tested from the Indian River Lagoon region had total mercury levels that were greater than or equal to the 0.5-ppm threshold level (.r=0.77 ppm; median=0.74 ppm; range 0.24- 1.7 ppm) (Table 1). There was no significant differ- ence between lengths of males (x =735 mm PCD and females (J =754.6 mm PCL; ^-test, P>0.5) nor was there a significant difference between mean total mercury levels between males and females (S^=0.76 ppm for both sexes; Mann-Whitney rank sum test, P>0.05). Ranges of total mercury for both sexes were also similar. There was a significant positive correlation be- tween total mercury level and bull shark length (both sexes combined) (P<0.001; Fig. 1). Total mercury lev- els in juvenile bull sharks increased as individuals grew larger, although some small sharks contained levels as high as those in larger sharks. Blacktip shark, Carcharhinus limbatus Total mercury levels for 21 juvenile and adult black- tip sharks examined ranged from 0.16 to 2.3 ppm 374 Fishery Bulletin 97(2), 1999 ln(Hg) = -1.57 + 0.0016 (PCL) I 5. S 1 « = 53 > t 0 u c -1 -" — V.«« • '/ • ••• . • • c -2 ■ ■ • ■ r ■ I 500 600 700 800 900 1.000 1.100 1.200 Precaudal length (mm) Figure 1 Relation between In total mercury level (ppm) and precaudal length (mm) for juvenile bull sharks. Carcharhinus leucas. from the Atlantic coast of Florida. ln(Hg) = -2.40 + 0.0018 (PCL) 2 r r = 0.538 c D. S 1 > 3 s -1 « = 21 c -2 400 600 800 1.000 1.200 1.400 1.600 1.800 Precaudal length (mm) Figure 2 Relation between In total mercury level (ppml and precaud al length (mm) for juvenile and adult blacktip sharks, Carcharhinus limbatiis. from the Atlantic coast of Florida. Table 1 Total mercury levels in sharks from the Atlantic coast of Florida, n = number of sharks analyzed. Total mercury Precaudal (ppm wet we ght) length mm) Standard Species Common name n Life history stage Range Mean deviation Range Mean Carcharhinus leucas bull shark 53 neonate/juvenile 0.24-1.7 0.77 0.32 552-1075 755.6 Carcharhinus limbatus blacktip shark 21 juvenile/adult 0.16-2.3 0.77 0.71 513-1.623 939.6 4 embryo 0.63-0.78 0.69 0.07 225-232 228.7 Rhizoprwnodon Atlantic sharpnose 81 juvenile/adult 0.11-2.3 1.06 0.63 220-857 591.6 terraenoiae shark 6 embryo 0.17-0.29 0.22 0.02 74-85 80.3 Sphyrna tiburo bonnethead shark 95 juvenile/adult 0.13-1.5 0.50 0.36 297-1081 590.3 41 embryo 0.08-0.35 0.16 0.07 206-255 232.1 (x=0.77 ppm; me(dian=0.44 ppm) (Table 1). Mean total mercury levels for females (x=0.87 ppm) were higher than those for males (.v=0.58 ppm): however, these (differences were not analyzed for statistical significance because sample sizes were small (« = 14 females; n=7 males). The female blacktip sharks col- lected in this study were larger (.r =994.3 mm PCL) than the males (.r=830.1 mm PCL). Almost half (47.6'^) of the blacktip sharks examined had mer- cury levels that were greater than or equal to the 0.5-ppm threshold level. The four blacktip shark embryos (x=235 mm PCL), collected from a single 1050-mm-PCL female whose total mercury level was 2.3 ppm, had total mercury levels ranging from 0.63 to 0.78 ppm (.r=0.69 ppm; median=0.68 ppm) (Table 2). Mercury levels in these embryos equaled 27.4-33.9% of levels observed in their mother. There was a significant positive correlation be- tween total mercury level and blacktip shark length (both sexes combined)(P<0.0001; Fig. 2). Total mer- cury levels in blacktip sharks increased as individu- als grew larger, although some small sharks con- tained levels as high as those in larger sharks. Atlantic sharpnose shark, Rhizoprionodon terraenovae Total mercury levels for 81 juvenile and adult Atlan- tic sharpnose sharks ranged from 0.11 to 2.3 ppm (.V = 1.06 ppm; median=0.95 ppm) (Table 1). Although NOTE Adams and McMichael: Mercury level in four species of sfiarks from tfie Atlantic coast 375 Table 2 Total mercury leve s of pregnant female sharks and associated embryos from the Atlantic coast of Florida. Pregnant female Embryos Precaudal length Total mercury Precaudal length Sex Total mercury Species (mm) (ppm) (mm) (ppm) Carchar-hinus Iimbatus 1050 2.3 230 227 235 223 M M M F 0.78 0.73 0.63 0.63 Rhizoprionodon terraenovae 766 1.9 84 83 F M 0.29 0.29 782 1.6 76 74 M M 0.17 0.18 784 2.3 85 80 F F 0.19 0.20 Sphyrna tibitro 990 1.1 229 233 236 M F F 0.12 0.12 0.14 855 1 225 237 227 M F F 0.13 0.19 0.12 1081 1.5 238 230 F M 0.35 0.14 Sphyrna tiburo 768 0.53 239 229 241 224 240 F F M F M 0.32 0.30 0.26 0.28 0.26 877 1.1 238 215 228 239 230 F M M F M 0.11 0.11 0.13 0.21 0.10 females (.v=624 mm PCL) were significantly larger than males (.v =541.9 mm PCL) (^test, P<0.05), to- tal mercury levels of males (.V =0.92 ppm) and females (3c =1.15 ppm) were not significantly different (Mann- Whitney rank sum test, P>0.1). A total of 72.8% of all juvenile and adult Atlantic sharpnose sharks tested had mercury levels that were greater than or equal to the 0.5-ppm threshold level. Total mercury levels for the six embryos (74-85 mm PCL) examined ranged from 0.17 to 0.29 ppm (x=0.22 ppm; median=0.19 ppm). Mercury levels for embryos within each litter were very similar (Table 2) and equaled 8.3-15.3'^ of levels observed in their respective mothers. Overall mercury levels among litters were also similar (±0.02 SD). All confirmed pregnant females had levels greater than 1.0 ppm. There was a significant positive correlation be- tween total mercury level and Atlantic sharpnose shark length (both sexes combined)(P<0.0001; Fig. 3). Total mercury level in this species increased as individu- als grew larger. Total mercury levels for larger At- In(Hg) = -: .35 + 0.0036 (PCL) -) r r = 0.697 « = 81 F D. D. 1 > ^^'-< ^ ^w • • 3 • ^A • . 1 E -1 .<••. • • ^ ••• • • a • ^^ • • c --■*• c -2 % 100 200 300 400 500 600 700 800 900 Precaudal length (mm) Figure 3 Relation between In total mercury- level (ppm) and precaudal length (mm) for juvenile and adult Atlantic sharpnose sharks, i?/!izopr;onorfon terraenovae, from the Atlantic coast of Florida. 376 Fishery Bulletin 97(2), 1999 lantic sharpnose sharks (>500 mm PCD were always greater than 0.5 ppm. Bonnethead shark, Sphyrna tiburo Total mercury levels for the 95 juvenile and adult bonnethead sharks sampled (297-1081 mm PCD ranged from 0.13 to 1.5 ppm (3c =0.50 ppm; me- dian=0.29 ppm) (Table 1). Forty percent of all juve- niles and adults tested had mercury levels that were greater than or equal to the 0.5-ppm threshold level. Mean lengths of males (.v=564.6 mm PCD and fe- males (x=619 mm PCL) were not significantly dif- ferent (Mann-Whitney rank sum test, P>0.1) and there were no significant differences in mercury lev- els between males (x=0.49 ppm) and females (x = 0.50 ppm) (Mann-Whitney rank sum test, P>0.5). There was a significant positive correlation between total mercury level and bonnethead shark length (both sexes combined) (P<0.0001; Fig. 4). Total mercury levels for the 41 embryos examined (206-255 mm PCL) ranged from 0.08 to 0.35 ppm (.r=0.16 ppm; median=0.13 ppm). Mean length for male and female embryos were both 232-mm PCL. Total mercury levels of male (x=0.15 ppm) and fe- male (x=0.18 ppm) embryos were similar (Table 2). Mercury levels for embryos within each litter were similar (+ 0.03 SDXTable 2), as were overall mer- cury levels among litters (mean mercury level/litter - 0.1-0.28 ppm). Total mercury levels in embryos of this species equaled 9.1-60.4% of levels observed in their respective mothers. In(Hg) - -2.69 + 0.0029 (PCLJ 2 - r = 0.779 n = 95 p 5. a- 1 p ^^ 'J 1— •• • = -1 • • o u, p .") *'• " 2( )0 400 600 800 1,000 1.200 F'rccaudal length ( mm 1 Figure 4 Relation between In total nierCury level (ppm) and precaudal length (mm) for juvenile and adult bonnethead sharks, Sphyrna tibiirn. from the At- lantic coast of Florida. Discussion Results of this study indicated that most individu- als in the four species of sharks tested had accumu- lated levels of total mercury that were as high as or higher than the regulatory threshold levels. Approxi- mately 60% of the juvenile and adult sharks tested in this study had total mercury levels that were greater than or equal to the 0.5-ppm threshold level, and 12% had levels greater than 1.5 ppm. Informa- tion regarding mercury in sharks from other areas is limited; however, elevated levels of mercury in various shark species have been documented in sev- eral regions, including waters off Florida and the southeastern United States (Gardner et al., 1975; Hueter et al., 1995; FHRS-), Canada (Forrester et al., 1972), Great Britain (Vas, 1991), and Austraha (Lyle, 1984, 1986). Although the life-history patterns of different shark species are diverse, most sharks grow slowly and have long lifespans. These general life-history characteristics, together with the high trophic status of many shark species, may contrib- ute significantly to the accumulation of high concen- trations of mercury (Lyle, 1984). Mercury levels were directly related to shark size in all species tested. Length differences accounted for 78%' of the total variation in mercury levels ob- served in bonnethead sharks, nearly 70% in Atlantic sharpnose sharks, and approximately 54% in blacktip sharks. The relation between mercury levels and length of juvenile bull sharks was less clear; length differences accounted for only 24% of the variation in mercury levels. The comparatively weak relation between mercury level and size in this species may stem from the fact that only neonates and small ju- veniles were sampled. Growth rates of bull sharks at the juvenile stage are thought to be quite variable (Dodrill, 1977;Thorson and Lacey 1982; Branstetter and Stiles, 1987), which may explain some of the observed variability of mercury levels in juveniles of this species. For species we examined, the relations between total mercury and shark size demonstrated that larger, presumably older, individuals accumu- lated higher levels of mercury. The accumulation of mercury with increasing size and age is likely to be a result of the slow and inefficient elimination of mercury from fish tissue in relation to the rate of accumulation (Rodgers and Beamish, 1982; Bryan, 1984). Because mercury levels increase as individuals grow larger, levels in juveniles can potentially be viewed as minima for the overall population. Al- though all bull sharks tested were either neonates or juveniles and the majority of blacktip sharks were juveniles or young adults, both species typically had NOTE Adams and McMichael: Mercury level in four species of sharks from tfie Atlantic coast 377 high mercury concentrations. High mean methyl- mercury levels (>1.0 ppm wet weight) were also de- tected in 29 larger size-class bull sharks (approx. 1300-2800 mm TL> and six blacktip sharks (> approx. 1500-1800 mm TL) collected off Florida (Hueter et al., 1995). Two of these bull sharks were collected off east-central Florida; however, specific size or sex data were not reported for these individuals. The observed relation of length to total mercury in juvenile bull sharks, together with results reported for larger size- class bull sharks and other closely related species (Lyle, 1984, 1986; Hueter et al., 1995), indicates that large adult bull sharks could contain excessive mer- cury levels. Blacktip sharks collected off northern Australia, which were of similar mean length and size range to those examined in our study, also had relatively high total mercury levels (Lyle, 1984, 1986). These data suggest that through bioaccumulation, larger, older adults of these common species poten- tially contain excessive mercury burdens. Total mer- cury levels for larger size-class Atlantic sharpnose sharks (>500 mm PCD that we examined were con- sistently greater than 0.5 ppm. Muscle tissue from the related Australian sharpnose shark, Rhizo- prionodon taylori. and the milk shark, R. acutus, tested from northern Australian waters (Lyle, 1986) contained mercury levels that were comparable to levels we detected in Atlantic sharpnose sharks. Milk sharks of comparable size examined by Lyle ( 1986 ) had total mercury concentrations (range=0. 16-2.00 ppm wet weight; .v=1.01 ppm) that were very similar to those in Atlantic sharpnose sharks in the current study. Bonnethead sharks had the lowest mean mercury content of the four species tested. The relatively low levels found in bonnethead sharks may be related to the comparatively fast growth rate and short life span of this species; however, diet is also an important factor. Bonnethead sharks off Cape Canaveral, Florida (senior author, unpubl. data), as well as off southwest Florida (Cortes et al., 1996), feed princi- pally on crustaceans, whereas the other three spe- cies consume a larger proportion offish (Compagno, 1984; Snelson et al., 1984; Castro, 1996; senior au- thor, unpubl. data). Although available data are lim- ited, crustaceans may contain mercury levels that are lower than those found in many fish species (Gardner et al.. 1975; Stickney et al., 1975; Jop et al., 1997). Mercury levels observed in embryos and neonates, as well as comparisons between pregnant females and associated embryos, indicated that transmission of mercury from maternal sources may be an impor- tant factor contributing to total mercury concentra- tions in shark muscle tissue. Total mercury levels in embryos of all species examined ranged from 8.3 to 60.4% of levels observed in their mothers, whose mercury levels were greater than the 0.5-ppm thresh- old (Table 2). Mercury levels of embryos within all litters were found to be similar. Mercury concentra- tions in embryos were variable among species. At- lantic sharpnose and bonnethead shark embryos all contained total mercury levels below the 0.5-ppm threshold; in contrast, all four embryos collected from a 1050-mm-PCL female blacktip shark had mercury levels greater than the threshold level. Variations in mercury concentrations found in embryos of differ- ent species could stem from differences in gestation periods or related physiological differences. Blacktip sharks have a two-year reproductive cycle with a 12- month gestation period (Castro, 1996), whereas At- lantic sharpnose sharks have a one-year reproduc- tive cycle with a 10—11 month gestation period (Par- sons, 1983). Bonnethead sharks have a gestation period of only 4-5 months (Parsons, 1993) — the short- est known gestation period of any placental vivipa- rous shark species (Parsons^). This short gestation period, combined with other physiological factors and maternal diet may explain why bonnethead shark embryos contained the lowest mean mercury levels in this study. Walker ( 1976) suggested that transfer of mercury to developing ova and embryos may re- duce mercury levels in mature females; however, little is currently known about maternal transmis- sion of mercury in fishes. Mercury levels were rela- tively high in large females despite transfer of mer- cury to developing ova and embryos. Results of this study indicate that the majority of sharks examined accumulated levels of total mercury that were greater than or equal to the threshold level of 0.5 ppm determined by the state of Florida. Mercury concentrations in adults and larger juveniles were t^-pi- cally elevated for all species; however, relatively high concentrations were also fi-equently observed in smaller juvenile bull and blacktip sharks. These data support the current health advisory in Florida urging limited consumption of sharks because of elevated mercury concentrations. Our data, in conjunction with the re- sults of other studies conducted in the region ( Gardner etal., 1975; Hueter etal., 1995; FHRS^), illustrate that mercury concentrations in several commonly landed shark species from the southeastern United States of- ten exceed state and federal regulatory levels. Acknowledgments We thank J. Arrecis, J. Brooks, W. Coppenger, and T. Fitzpatrick of FDEP Division of Technical Services ^ Parsons, G. R. 1997. Department of Biology, University of Mississippi, University. MS 38677. Personal commun. 378 Fishery Bulletin 97(2), 1999 for laboratory analyses that made this study possible. We greatly appreciate the efforts of commercial gillnet and longline fishermen in the Cape Canaveral region who allowed us to sample sharks at sea or dockside, especially A. Christopher, who frequently allowed us onboard his vessel to collect sharks. We also thank the fisheries crew and port samplers (in particular, M. Ridler) of the FDEP Indian River Field Laboratory for assistance in collecting and process- ing samples and J. Guenthner for shipping and pro- cessing samples. Cape Canaveral Scientific, Inc. also provided sharks for this study. Lastly, we thank L. French, J. Leiby, G. Parsons, K. Peters, J. Quinn, T. Sminkey, and D. Tremain who offered helpful sug- gestions for improving the manuscript. Literature cited Barber, R. T., and P. J. Whaling. 1983. Mercury in marlin and sailfish. Mar. Pollut. Bull. 14:395-396. Bloom, N. S. 1992. On the chemical form of mercury in edible fish and marine invertebrate tissue. Can. J. Fish. Aquat. Sci. 49:1010-1017. Branstetter, S., and R. Stiles. 1987. Age and growth estimates of the bull shark, Carcha- rhinus leucas. from the northern Gulf of Mexico. Environ. Biol. Fishes 20(31:169-181. Brown, S. T. In press. Preregulatory developments in the Florida com- mercial shark fishing industry with implications for the recreational fishery. North Am. J. Fish. Manage. Bryan, G. W. 1984. Pollution due to heavy metals and their compounds. In O. Kinne (ed.i. Marine ecology, vol. 5, p. 1289-1431. Wiley, London. 1993. The shark nursery of Bulls Bay. South Carolina, with a review of the shark nurseries of the southeastern coast of the United States. Environ. Biol. Fishes 38:37-48. Castro, J. I. 1996. Biology of the blacktip shark, Carcharlunun limbatus, off the southeastern United States. Bull. Mar. .Sci. 59(3):.508-.522. Compagno. L. J. V. 1984. Sharks of the world: an annotated and illustrated catalogue of shark species known to date. FAG .Species Catalogue, vol. 4, parts 1 and 2. FAQ Fish. Synop. 125, FAG, Rome, 655 p. Cortes, E., Manire, C. A., and Hueter, R. E. 1996. Diet, feeding habits, and diel feeding chronology of the bonnethead shark, Sphyrna tiburo. in southwest Florida. Bull. Mar .Sci. 58:35.3-367. Dodrill, J. W. 1977. A hook and line survey of the sharks found within five hundred meters of shore along Melbourne Beach. Brevard County. Florida. M..S. thesis, Florida Inst. Technol,. Melbourne, Fl„ 304 p. EPA (U.S. Environmental Protection Agency). 1991. Determination of mercury in tissues by cold vapor atomic absorption spectrometry: method 245.6 (revision 2.31. U.S. Environmental Protection Agency, Environmen- tal Monitoring Systems Laboratory, Cincinnati, GH, 13 p. Forrester, C. R., K. S. Ketchen, and C. C. Wong. 1972. Mercury content of spiny dogfish, Squalus acanthias, in the Strait of Georgia, British Columbia. Can. J. Fish. Aquat. Sci. 29(10):1487-1490. Forstner, U., and G. T. W. Wittman. 1981. Metal pollution in the aquatic environment. 2nd ed. Springer- Verlag, Berlin, 486 p. Fox, E., J. Kuo, L. Tilling, and C. Ulrich. 1994. SigmaStat user's manual, rev. ed. SSW 1.0. Jandel Scientific Corp.. San Rafael, CA, 656 p. Gardner, W. S., H. L. Windom, J. A. Stephens, F. E. Taylor, and R. R. Stickney. 1975. Concentrations of total mercury and methylmercury in fish and other coastal organisms: Implications of mer- cury cycling. In F. G. Howell, J. B. Gentry, and M. H. Smith (eds.). Mineral cycling in southeastern ecosystems, p. 268-278. ERDA Symposium Series, Springfield. VA. [Available from National Technical Information Service (CONF-7405131.J Grieb, T. M., C. T. Driscoll, S. P. Gloss, C. L. Schofield, G. L. Bowie, and D. B. Porcella. 1990. Factors affecting mercury accumulation in fish in the upper Michigan peninsula. Environ. Toxicol. Chem. 9:919-930. Hueter, R. E., W. G. Fong, G. Henderson, M. F. French, and C. A. Manire. 1995. Methylmercury concentration in shark muscle by species, size and distribution of sharks in Florida coastal waters. Water Air Soil Pollut. 80:893-899. Jop, K. M., R. C. Biever, J. L. Hoberg, and S. P. Shepard. 1997. Analysis of metals in blue crabs, Callinecles sapidus, from two Connecticut estuaries. Bull. Environ. Contam. Toxicol. 58:311-317. Kai, N., T. Ueda, Y. Takeda, and A. Kataoka. 1987. Accumulation of mercury and selenium in blue marlin. Nippon Suisan Gakkaishi. 53f9):1697, Lyie, J. M. 1984. Mercury concentrations in fourcarcharhinid and three hammerhead sharks from coastal waters of the Northern Territoiy Aust. J. Mar Freshwater Res. 35:441-51. 1986. Mercury and selenium concentrations in sharks from northern Australian waters. Aust. J. Mar Freshwater Res. 37:309-21. Meaburn, G. M. 1978. Heavy metal contamination of Spanish mackerel, Scomberomorus maculatiis, and king mackerel, .S'. caualla. In Proceedings of the mackerel colloquium, March 16, 1978, p, 61-66. Charleston. Lab . SEF.SC. Charleston. .SC. Miller, G. E., P. M. Grant, R. Kishore, F. J. Steinkruger, F. S. Rowland, and V. P. Guinn. 1972. Mercury concentrations in museum specimens of tuna and swordfish. Science (Wash., D.C.) 175:1121-1122. NMFS (National Marine Fisheries Service). 1993. Federal management plan for sharks of the Atlantic Ocean. U.S. Dep. Commer, Natl. Mar Fish. -Serv., South- east Regional Office. St. Petersburg, FL, 167 p. Parsons, G. R. 1983. The reproductive biology of the Atlantic sharpnose shark. Rhizopnimodon terraenovae . Fish. Bull. 81:61-73. 199.3. Geographic variation in reproduction between two popu- lations of the bonnethead shark, Sphyrna tiburo. Environ. Biol. Fishes 38:2.5-35, Phillips, G. R., and D. R. Buhler. 1978. The relative contributions of methylmercury from NOTE Adams and McMichael: Mercury level in four species of shiarks from the Atlantic coast 379 food or water to rainbow trout, Salmo gairdneri. in a con- trolled laboratory environment. Trans. Am. Fish. Soc. 107161:853-861. Rodgers, D. W., and F. W. H. Beamish. 1982. Dynamics of dietary methylmercury in rainbow trout, Salmo gairdnen. Aquat. Toxicol. 2:271-290. Snelson, F. F. Jr., T. J. Mulligan, and S. E. Williams. 1984. Food habits, occurrence, and population structure of the bull shark, Carcharhinus leiicas. in Florida coastal lagoons. Bull. Mar Sci, 34:71-80. Sokal, R. R., and F. J. Rohlf. 1981. Biometry Freeman, New York, NY, 8.59 p. Stickney, R. R., H. L. Windom, D. B. White, and F. E. Taylor. 1975. Heavy metal concentrations in selected Georgia es- tuarine organisms and comparative food habit data. In F. G. Howell, .J. B. Gentry, and M. H. Smith (eds.). Mineral cycling in southeastern ecosystems, p. 256-267. ERDA Symposium Series, Springfield, VA. [Available from Na- tional Technical Information Service (CONF-740513).l Thorson, T. B., and E. J. Lacey Jr. 1982. Age, growth rate and longevity of Carcharhinus leucas estimated from tagging and vertebral rings. Copeia 1982(1):110-116. Trent, L., D. E. Parshley, and J. K. Carlson. 1997. Catch and bycatch in the shark driftnet fishery off Georgia and east Florida. Mar Fish. Rev 59(1): 19-28. Turner, M. D., D. O. Marsh, J. C. Smith, J. B. Inglis, T. W. Clarkson, C. E. Rubio, J. Chiriboga, and C. C. Chiriboga. 1980. Methylmercury in populations eating large quantities of marine fish. Arch. Environ. Health 35(6):367-378. Vas, P. 1991. Trace metals in sharks from British and Atlantic waters. Mar Pollut. Bull. 22(2):67-72. Walker, T. I. 1976. Effects of species, sex, length and locality on the mer- cury content of school shark, Galeorhinus australis. and gummy shark, Mustelus antarcticus, from South-eastern Australian waters. Aust. J. Mar Freshwater Res. 27:603-616. 380 A unique shell marker In juvenile, hatchery-reared individuals of the softshell clam, Mya arenaria L. Brian F. Beal University of Maine at Mactiias 9 O'Brien Avenue, Machias, Maine 04654 E-mail address bbeal aacad umm maine edu Robert Bayer University of Maine Hitchner tHall, Orono, Maine 04469 M. Gayle Kraus University of Maine at Machias 9 O'Brien Avenue, Mactiias, Maine 04654 Samuel R. Chapman University of Maine Darling Marine Center, Walpole, Maine 04573 The ability to identify individual bivalve mollusks in field popula- tions is fundamental to under- standing potential population regu- latory mechanisms (such as the in- fluence of population density, tidal height, and initial shell size on growth and survival rates, and fe- cundity schedules). Softshell clams, Mya arenaria L., have been com- mercially harvested from the soft- bottom intertidal zone in Maine since the mid-1800s and form the basis of an extensive fishery along its entire coast. Dramatic declines in landings during the past decade in Maine (Wallace, 1997 ), however, have resulted in attempts to use hatchery-reared juveniles to supple- ment wild stocks (Beal, 1994). In the past, distinguishing between cul- tured and wild bivalves in the field in order to follow their fate has been performed by using alizarin stain- ing techniques (Newell and Hidu, 1982), by tagging individuals (Brousseau, 1979), or by applying colored marks (e.g. paint dots) to the valves (Peterson and Beal, 1989). Here, we describe a natural and unique shell marker for juve- nile, hatchery-reared softshell clams that forms on the outer valves of individuals once they are placed in the field. The distinctive mark appearing on the surface of each valve obviates the need to ap- ply physical tags to individuals and also eliminates the stress that small clams otherwise undergo when being tagged. Methods Clams were reared during the sum- mer of 1983 at a 4-H clam hatch- ery in Jonesboro, Maine, and were held in running seawater in sedi- ment-free trays at the Darling Ma- rine Center, Walpole, Maine, until 25 May 1984. On that date, 5000 individuals (mean shell length [x^^ ±1 SD1 = 10.4 ±1.52 mm; range= from 6.7 to 14.6 mm; n=200) were transported back to the Jonesboro hatchery where the clams were di- vided into five groups of 100, 200, 300, and 400 individuals (i.e. a to- tal of 20 groups ). All but ten individu- als from each group (20 groupsx 10 individuals=200 clams) were marked with a single, group-spe- cific, color-coded dot (Mark-Tex Corp. paint) on both valves near the umbo. The remaining 200 clams were painted with two dots to en- sure individual recognition and measured (greatest shell length) to the nearest 0.1 mm by using ver- nier calipers. Marking was per- formed at this time because it was not known whether a distinctive mark would form naturally on the surface of each valve once clams began adding new shell in the field. Clams were maintained overnight without seawater at the University of Maine at Machias in a walk-in cooler (ca. 5°C). On 25 May 1984, a matrix con- sisting of two rows of ten 0.25-m^ plots was established near the midtide level of an intertidal fiat (the Narrows) located along the western shore of the Chandler River (44°39'04"N; 67°33'10"W) near the town of Jonesboro, Maine. Sediments consisted of poorly sorted muds with a graphic mean ±1 SD of 3.6 ±0.34(t) (N=2). The ma- trix was located approximately 55 m from the extreme high water mark and 35 m from the mean low water mark. Both rows were par- allel to the shore and were spaced 1 m apart. Black rigid-mesh enclo- sures (DUPONT Vexar, 6.4 mm aperture), approximately 0.25 m^ and 15-cm wide, were installed around each plot to restrict lateral movements of the small clams (Baptist, 1955). Because the enclo- sures had no mesh roofs, they al- lowed epibenthic and infaunal predators isensu Commito and Ambrose, 1985) access to the clams (Peterson and Beal, 1989). The mesh of each enclosure wall was Manuscript accepted ."j May 1998. Fish. Bull. 97;380-.386 (1999). NOTE Beal et al : Shell marker for juvenile, hatchery-reared Mya arenaria 381 attached by staples to a 40 x 2.5 x 2.5 cm wooden stake at each corner. This square, fenced assembly was then pushed 8 to 10 cm into the sediments leav- ing a mesh wall projecting from 5 to 7 cm vertically. Groups of clams were assigned randomly to the open enclosures on 26 May 1984. This completely random design resulted in five replicates of each of four evenly spaced planting densities ranging from 400 to leOO/m-. The fate of the 200 clams marked with two dots is reported here (see Beal [1994] for results of increasing intraspecific density on the fate and growth of the remaining clams). To ensure that seagulls (Lams spp.) did not prey on the clams be- fore they had the opportunity to burrow, one of us remained at the site until the tide had completely covered the clams and the enclosures. During this interval (ca. 2 hours) clams burrowed into the sedi- ments approximately 20 minutes after planting. No clam was visible at the sediment surface at the time of tidal inundation. Live and dead clams were recovered from each of the twenty open enclosures after 99 days ( 1 Septem- ber 1984) by collecting the top 15 cm of sediment from each enclosure and sieving the sediment through a 0.5-mm mesh. For each live clam (marked with two dots or not), an obvious mark parallel to the entire shell circumference was visible on the shell at or near the size the animal was on 26 May 1984 (see "Re- sults" section). This mark permitted us to estimate a shell growth index for every live clam: Relative growth index = [{final length - initial length) / initial length] x 100. Index values greater than 100% indicate at least a doubling in growth. Shells from this 1984 field experiment were not saved, but results from subsequent field trials in eastern Maine with hatchery-reared softshell clam juveniles (Beal, 1994) revealed that this shell mark remains visible for periods up to 1.5 years after plant- ing. We used scanning electron microscopy (SEM) to describe the gross shell structure across the distinc- tive pre- and postplanting regions of the outer shell of juvenile, hatchery-reared clams. We used the valves of cultured clams (initial size similar to those reported above) that had been reared during 1989 at the Beals Island Regional Shellfish Hatch- ery (BIRSH), Beals, Maine, planted at the Narrows in May 1990 and recovered in October 1990. We also used SEM to examine the outer shell struc- ture of wild juvenile clams (ca. 15 mm) taken from a nearby mudflat in Roque Bluffs, Maine, in Sep- tember 1992. Results When viewed macroscopically, every live hatchery- reared clam taken on 1 September 1984 (regardless whether or not it had been marked with two dots) had a distinct line on its shell surface that appeared to correspond to its size at planting on 26 May 1984. Scanning electron microscopy (SEM) of the gross shell structure of wild juvenile clams (Fig. 1, A and B) revealed sharp ridges (=lines viewed macroscopi- cally) approximately 15-microns apart punctuated by a relatively smooth region. In sharp contrast, the valves of cultured clams (Fig. 1. C and D) consistently revealed three distinct zones: an inner region from the umbo to a distinct area (line) of demarcation that was approximately 50-microns in width (Fig. ID) and an outer zone from that line to the ventral margin. The inner zone was characterized by a pitted, amor- phous surface with interrupted ridges, whereas the outer zone of the cultured clams (Fig. IC) appeared similar to the entire surface of wild clams. Observa- tions from experimental or stock enhancement outplantings of cultured clams from BIRSH in nu- merous coastal communities in Maine since 1987 indicate that the shell mark may be present for up to 1.5 years after planting (Beal, 1994). To test whether the obvious line, or shell mark, could be used quantitatively to distinguish initial size at planting from subsequent shell growth, we exam- ined closely the 57 surviving clams of the 200 that had been marked with two dots on 25 May 1984. We used calipers to estimate the initial size of the clam delimited by the obvious surficial shell mark (line). Next, we compared, for each clam, this estimate of initial (i.e. predicted) shell length with the one re- corded on 25 May. If the hypothesis that the line is formed at or near the time of transplanting is cor- rect, then the mean difference between actual (i.e. recorded) and estimated size should not be signifi- cantly different from zero. Results of a one-sample, two-tailed /-test demonstrated that the average dif- ference was not significantly different from zero (- 0.047 mm; ^=-1.505; P=0.138; n=57). An alternative statistical approach is to plot the predicted initial shell length (y-axis) against the actual initial shell length (.v-axis; Fig. 2) and to determine whether the straight line (y= -0.229 -f- 1.017.v; r~=0.97) is signifi- cantly different from the line y = x. The slope of the least squares equation for that line was not signifi- cantly different from unity (F=1.38; df=2, 55; P=0.26), and the intercept was not significantly different from zero (F=0.81; df=l, 55; P=0.37). The mean relative growth index for the period 26 May to 1 September for the 57 marked survivors was 86.9 ± 2.26% SE (Fig. 3A). Mean final length of the same in- 382 Fishery Bulletin 97(2), 1999 10KV X17 1000U 001 06082 UM Figure 1 Scanning electron microscopy (SEM) of the shell surface of a wild (A and B) and cultured (C and D) juvenile softshell clam. Mya arenana (length ca. 15 mm). The wild clam was sampled from a mudflat in Roque Bluffs, Maine, in September 1992. The cultured clam was reared at the Beals Island Regional Shellfish Hatchery and planted in May 1990, in a mudflat at the mouth of the Chandler River, near the town of Jonesboro, Maine. Photo B, taken near the umbo region of the wild clam, shows uninterrupted concentric ridges and a lack of pits and amorphous grooves. Photo C is an example of "the hatchery mark." It is a continuous, ca. 50-micron groove that extends along the ventral slope (anterior to posterior) of both valves and appears at that point on the shell surface associated with the time the animal leaves the hatchery and is placed in the soft-bottom benthos. Photo D, was taken along the anterior-posterior groove that appears at the interface before the period when new shell is deposited once the individual is trans- planted to the field. The white bar indicates a length of 1000 n for A and C and 100 \i for B and D. NOTE Beal et al : Shell marker for juvenile, hatchery-reared Mya arenaria 383 0KV X18 10e0U 088 06082 UM Figure 1 (continued) dividuals was 19.6 ±0.24 mm SE (Fig. 3B), which equates to an average gi-owth rate of 0.093 mm/day. Discussion Our data provide evidence of a distinctive mark or Une being formed on the outer surface of the valves of juvenile hatchery-reared softshell clams. We have observed from other field tests and stock enhance- ment trials in Maine that the mark may be present for up to 1.5 years after planting, after which, the line disappears as the umbo region of the shell be- comes increasingly worn, presumably on account of sediment abrasion from burrowing or seasonal repo- sitioning (Zwarts and Wanink, 1989). The formation of the line on the shell surface be- tween the umbo and the ventral margin allows one to distinguish easily between wild and cultured ju- venile clams and can be a valuable tool for resource 384 Fishery Bulletin 97(2), 1999 14-1 ed initial shell length 1 1 y 6 8 10 12 14 Actual initial shell length (mm) Figure 2 Predicted initial shell size versus actual initial shell size for the 57 hatchery-reared individuals of Mya arenaria marked w^ith two dots that survived in open 0.25-m- field enclosures during the period from 26 May to 1 September 1984. managers wishing to assess the fate of hatchery seed for stock enhancement purposes (Beal, 1994). In ad- dition, it obviates the need to mark (physically or chemically) individuals or a group. If initially un- marked, juvenile, hatchery-reared softshell clams are planted in the field and examined within an 18- month period, the distinctive shell mark allows one 1) to determine initial planting size of individual hatchery-reared clams (Fig. 2); 2) to determine rela- tive or absolute growth rates (Fig. 3, A and B); and 3) to estimate the time (e.g. season) when death oc- curred for those individuals that suffered mortality after planting. Similar disturbance lines were ob- served on juvenile, hatchery-reared northern qua- hogs, Mercenaria mercenaria (L.), planted from the laboratory to various subtidal and intertidal sites in North Carolina (Beal, 1983). The origin of the shell mark may be related to con- ditions within the hatchery environment. After the field experiment described above, cultured softshell clam juveniles from BIRSH were planted in inter- tidal flats along the coast of Maine ( 1987-94) during the months of April through October for both ma- nipulative tests and community-based stock enhance- ment efforts (Beal, 1994). In addition, cultured clam stock produced in a commercial shellfish hatchery (Mook Sea Farm, Inc., Walpole, Maine) were used in stock enhancement programs in Gloucester and Ipswich, Massachusetts (Whitten'). Regardless of 0 20 40 60 80 100 120 140 160 Relative grovirth index (x lOO) 30- 25- 20- 15 10- B I—I Initial longths ■ Final lengths Jill Ik r-" 10 15 20 25 Shell length (mm) 30 Figure 3 (A) Frequency distribution of relative growth rate index for the 57 hatchery-reared softshell clams marked with two dots (26 May to 1 September 1984). An index of lOO^-f indi- cates a doubling of shell length. iBl Frequency distribu- tion of initial (26 May 1984; n=200) and final shell lengths il September 1984; n=57) of hatchery-reared softshell clams. ' Whitten, J. 1996. Merrimak Valley Planning Commission, Haverhill, MA. Personal commun. hatchery origin, the same line appeared on the shells of all survivors planted in the soft-bottom intertidal zone at all locations, as well as on shells of those that grew before dying. The mark has been observed in clams planted at shell sizes as small as 3 mm and as large as 25 mm (Beal, 1994). In addition, this mark appears regard- less whether clams are transplanted to the field di- rectly from the hatchery or are overwintered in sub- merged, floating cages {sensu Beal et al., 1995) be- fore transplanting to the field. NOTE Beal et al,: Shell marker for luvenile, hatchery-reared Mya arenana 385 Although not specifically tested, at least two com- peting hypotheses may explain the mechanism that creates this shell marker. The first hypothesis is distur- bance, for clams in the hatchery are grown in sediment- free trays with fine-mesh screening (150-1500 \.i), handled (sieved) weekly, fed tropical species of micro- algae such as Isochrysis galbana (Tahitian variety), and are grown at temperatures between 20 and 25°C — well above those normally experienced in the wild in eastern Maine. Once they have left the hatch- ery environment, clams are planted on flats and im- mediately burrow into sediments where 1 ) hatchery- produced disturbance ends, 2) the clams begin feed- ing on natural phytoplankton assemblages, and 3) they experience seawater temperatures that typically do not exceed 17°C. The new shell that grows is al- ways distinctly whiter than the older shell. This dif- ference in coloration may indicate a release from com- petition for calcium or aragonite which may be lim- ited under hatchery conditions (Barber-). The sec- ond hypothesis is bacterial damage that may also relate to conditions in the hatchery where elevated levels of marine bacteria such as Vibrio spp. fre- quently occur. Qualitative tests for the presence of Vibrio spp. (using Difco TCBS agar [Elston et al., 1981] ) at BIRSH are made regularly and show a gen- eral presence of these bacteria in the seawater within tanks holding juvenile clams, in the cultured algae, and on the valves of juvenile clams. These gi'am-nega- tive, oxidase-positive, fermentive rods have been observed similarly in commercial bivalve hatcheries in the coastal northeastern United States (Elston, 1984) where they have been described as coating the shell surface of cultured juvenile northern quahogs, oysters, Crassostrea vi>-ginica (Gmelin), Ostrea editlis L., and bay scallops, Ar^opec^e/! irradians (Lamarck). Bacteria in the family Vibrionaceae can erode and perforate areas on the surface of bivalve shells through the production of a variety of acidic metabo- lites that are inimical to normal deposition of cal- cium carbonate (Elston et al., 1982). The SEM pho- tograph of the valve of the hatchery-reared softshell clam ( Fig. 1 , C and D ) clearly shows pitting and amor- phous grooves that may indicate a bacterial origin. During the past decade, clam landings in eastern Maine have declined by nearly 75^r (Wallace, 1997). Communities that manage their clam stocks in this and other regions along the coast are beginning to use cultured softshell clams to enhance clam produc- tion. Testing the biological and economic efficacy of hatch-and-release programs is critical for the devel- opment of sensible management programs. Results ^Barber, B. 1994. Universityof Maine. Orono, ME. Personal commun. presented here demonstrate the ease of distinguish- ing cultured from wild Mya and will allow scientists as well as clam harvesters a rapid assessment of field planting programs. Acknowledgments Support for this work was provided by a Maine/New Hampshire Sea Grant research grant to B. F. Beal and M. G. Kraus (R/FMD-191; NA89AA-D-SG020). Additional support was provided by the Department of Animal, Veterinary, and Aquatic Sciences at the University of Maine and by the University of Maine at Machias. We thank all those that helped in the 4-H hatchery and in the field, including M. Alley, A. Chapman, J. Chapman, J. Cox Jr., J. Cox Sr., D. Engels, C. Gay R. Look, S. Renshaw, and T. White. We also thank W. Congleton, S. Fegley, R. Hawes, A. Lewis, R. Vadas Sr., K. Vencile, and three anonymous reviewers for their useful comments on previous drafts of this note. Literature cited Baptist, J. P. 1955. Burrowing ability of juvenile clams. U.S. Fish Wild. Ser\'. Spec. Sci. Rep. 140:1-13. Beal, B. F. 1983. Effects of environment, intraspecific density, preda- tion by snapping shrimp and other consumers on the popu- lation biology of Mercenaria mercenaria near Beaufort, North Carolina. M.S. thesis, Univ. North Carolina, Chapel Hill.NC, 180 p. 1994. Biotic and abiotic factors influencing growth and sur- vival of wild and cultured soft-shell clams, Mya arenaria L., in eastern Maine. Ph.D. diss., Univ. Maine, Orono, ME, 499 p. Beal, B. F., C. D. Lithgow, D. P. Shaw, S. Renshaw, and D. Ouellette. 1995. Overwintering hatchery-reared individuals of the soft -shell clam, Mya arenaria L.: a field test of site, clam size, and intraspecific density. Aquaculture 130:14.5-158. Brousseau, D. J. 1979. Analysis of growth rate in Mya arenaria using the von Bertalanffy equation. Mar. Biol. 51:221-227. Commito, J. A., and W. G. Ambrose Jr. 1985. Predatory infauna and trophic complexity in soft- bottom communities. In P. E. Gibb (ed.). Proceedings of the nineteenth European marine biology symposium, p. 323-333. Cambridge Univ. Press, Cambridge. England. Elston, R. 1984. Prevention and management of infectious diseases in intensive mollusc husbandry. J. World Marie. Soc. 1.5:284-300. Elston, R., E. L. Elliot, and R. R. Colwell. 1982. Conchiolin infection and surface coating Vibrio: shell fragility, growth depression and mortalities in cultured oysters and clams, Crassostrea virginica. Ostrea edulis. and Mercenaria mercenaria. J. Fish Diseases 5:265-284. 386 Fishery Bulletin 97(2), 1999 Elston, R., L. Leibovitz, D. Relyea, and J. Zatila. 1981. Diagnosis of vibriosis in a commercial oyster hatch- ery epizootic: diagnosis tools and management features. Aquaculture 24:53-62. Newell, C. R., and H. Hidu. 1982. The effects of sediment type on growth rate and shell allometry in the soft-shell clam Mya arenaria L. J. Exp. Mar. Biol. Ecol., 65:285-295. Peterson, C. H., and B. F. Beal. 1989. Bivalve growth and higher order interactions: impor- tance of density, site, and time. Ecology 70:1390-1404. Wallace, D. E. 1997. The molluscan fisheries of Maine. In C. L. Mac- Kenzie Jr.. V. G. Burrell Jr., A. Rosenfield, and W. L. Hobart (eds.). The history, present condition, and future of the molluscan fisheries of North and Central America and Europe, vol. 1, Atlantic and Gulf coast, p. 63-86. U.S. Dep. Commer., NOAA Tech. Rep. 127. Zwarts, L., and J. Wanink. 1989. Siphon size and burying depth in deposit- and sus- pension-feeding benthic bivalves. Mar Biol. 100:227-240. 387 Occurrence of neonate and juvenile sandbar sharks, Carcharhinus plumbeus, in the northeastern Gulf of Mexico John K. Carlson Southeast Fistieries Science Center National Manne Fisheries Service, NOAA 3500 Delwood Beach Road, Panama City, Flonda 32408 E-mail address carlsoniffbiofsuedu The sandbar shark, Carcharhinus plumbeus, is a large, coastal spe- cies of the western north Atlantic occurring from Cape Cod, Massa- chusetts, to Brazil, including the Gulf of Mexico and Caribbean (Bigelow and Schroeder, 1948; Springer, 1960). Sandbar sharks are targeted by commercial fisher- ies and account for up to 60*^ of the large coastal shark landings in U.S. southern waters (NMFS, 1993). Adults are highly migratory and mostly congregate offshore. Neo- nate and juvenile sandbar sharks, on the other hand, are commonly found in coastal nursery areas where they feed ( Med vedetal., 1985) and avoid predation ( Springer, 1967; Branstetter, 1990) during summer months. The presence of neonate and ju- venile sandbar sharks has been well documented in coastal areas of the eastern United States. Springer (1960) reported juvenile sandbar sharks from Cape Cod, Massachu- setts, to Cape Canaveral. Florida. Juvenile sandbar sharks are abun- dant in Chesapeake Bay and east- ern shore of Virginia in less than 10 m (Medved and Marshall, 1981; Musicketal., 1993). Castro (1993) found neonate and juvenile sharks in Bulls Bay, South Carolina. Pratt and Merson' determined Delaware Bay, New Jersey, as a major nurs- ery area for neonate and juvenile sharks. Further attempts to delin- eate the extent of nursery areas for sandbar sharks along the U.S. east coast continue (Damon, 1997; Pratt and Merson'). Juveniles are not known to occur in coastal areas of the eastern Gulf of Mexico. The direct correlation of juvenile sandbar shark survivor- ship and future stock size (Cortes, in press) requires delineation of sandbar shark nursery areas. Thus, if recruitment to the stock is as- sumed to be entirely from sharks from the U.S. east coast, then un- derestimates of total population size could occur and affect overall stock assessments. This paper re- ports on the occurrence of neonate and juvenile sandbar sharks and the potential nursery area of these sharks in coastal waters of the northeastern Gulf of Mexico. Materials and methods Sandbar sharks were captured from October 1992 to October 1997 as part of studies on the distribution and abundance of sharks in the eastern Gulf of Mexico. Because sampling had various objectives, the variabil- ity in sampling design and methods precluded quantification of a valid time series of abundance (e.g. CPUE) from 1992 to 1997. In general, gill nets varied in height from 1.52 to 3.04 m and ranged in length from 30.4 to 273.6 m, and mesh sizes from 6.9 to 20.3 cm stretched mesh. Each net, regardless of size, was anchored at both ends and fished on the bot- tom. Longlines, which ranged in length from 76 to 335 m and con- sisted of 10-60 hooks, were anchored at both ends and fished so that one half fished the "midwater" and the other the "bottom." Gangions were 0.9-1.8 m long and hooks were size 3/0 and 12/0 ( Mustad). Usually men- haden (Brevoortia spp. ) was the bait of choice. The nets or longlines, or both, were set over a 24-h period at various times. Gill nets and longlines were checked, or checked and pulled, and sharks were removed throughout each sampling period. Surface water temperature (°C), salinity (ppt), and light transmission (cm) were mea- sured daily at each station. After they were caught, sharks were sexed and measured (total length, TL) to the nearest mm. Sharks that were in poor condition were euthanized for life history in- formation; those in good condition were tagged with a multirecapture, nylon-head, dart tag (Hueter and Manire'-) and released. Sampling took place April to October of each year, occasionally from November to March. Study area Sampling sites were located in four major areas along the northeastern portion of the Gulf Apalachee Bay to St. Andrews Bay, Florida (Fig. 1). 1 Pratt. H. L.. and R. R. Merson. 1996. Delaware Bay sandbar shark nursery pi- lot study. Narragansett Laboratory, Na- tional Marine Fisheries Service. NOAA. Progress report, 23 p. - Hueter, R. E., and C. A. Manire. 1994. By-catch and catch-release mortality of small sharks and associated fishes in the estuarine nursery grounds of Tampa Bay and Charlotte Harbor. NOAA NMFS/ MARFIN Program. Project Rep. NA17FF- 0378-01, 183 p. Manuscript accepted 13 May 1998. Fish. Bull. 97:387-391 (1999). 388 Fishery Bulletin 97(2), 1999 Apalachee Bay Indian Pass St. Vincent Island Florida Study Area Figure 1 Map of the study area in northwest Florida near latitude 30"00'N and longitude 85'35'W illus- trating the major sampling areas (o) and the two areas (•) in which sandbar sharks were captured from October 1992 to October 1997. The eastern part of this area has an irregular coast- line, few beaches and enclosed bay systems, and has large amounts of submergent (Thalassia spp. and Halodule spp.) and emergent vegetation (Spartina spp. and Juncus spp I. The western part has numer- ous barrier islands and sand beaches and is composed of semi-enclosed bays. Tidal amplitude in the bays is highest in Apalachee Bay and generally decreases toward the west. St. Andrew Bay consists of several embayments (average range 1.9-5.7 m deep) and has low fresh- water inflow, low turbidities, and high percentages of sand in the substrate. Salinity ranges from 13 to 32 ppt and tidal amplitude averages 0.48 m. The sys- tem exchanges water with the Gulf of Mexico through two passes, a natural pass at the east end and a man- made pass at the west end. St. Andrew Sound is a small semi-enclosed marine lagoon with expanses of submergent vegetation. It is about 14.5 km long and 0.2-2.0 km wide and has water depths from 3.5 to 4.5 m deep (mean high tide). Salinity ranges from 25 to 36 ppt and tidal ampli- tude averages 0.42 m. The sound exchanges water with the Gulf of Mexico through a pass (=0.5-2.0 km wide) that was created near the center of Crooked Island by Hurricane Eloise in 1975. Indian Pass is located at the western end of the Apalachicola Bay system. This area is about 2-3 km south of St. Vincent Island in the Gulf of Mexico where the average range of water depth is 5—10 m. The bay system surrounding this area is largely a line of barrier islands fronting the intersection of the Apalachicola delta and is the only bay system in Florida in which a large river system drains. As a result of river discharge, there is little submergent vegetation due to high turbidity. Salinity fluctuates from 15 to 35 ppt and tidal range is 0.66 m. Apalachee Bay is an open ocean bay without bar- rier islands separating the area from the open Gulf of Mexico. The bay is broad, shallow (average 3 m), and extends about 15 km offshore. Salinity ranges from 22 to 36 ppt and tidal amplitude averages 1.0 m. Wave energy is low and the area has large expanses of submerged vegetation. NOTE Carlson: Occurrence of neonate and luvenile Charchannus plumbeus in the northeastern Gulf of Mexico 389 Results Neonate and juvenile sandbar sharks (n = 105) were captured in two areas of the northeastern Gulf of Mexico, Indian Pass, and St. Andrew Sound. Captured sandbar sharks were 572-1640 mm TL (P^ig. 2). Seventeen were determined to be neonates and young-of-the-year (mean size=643 ±44.5 mm TL), as indi- cated by the presence of an open or par- tially healed umbilical scar. The larg- est sandbar shark with a partially healed umbilical scar was 720 mm TL. Following the age-length relationship provided in Sminkey and Musick ( 1995 ) and the life history stage classification in Cortes (in press) we estimated that 52 sharks were small juveniles ( about 1-3 yr old), 34 were large juveniles (about 4- 9 yr old), and 2 were subadults (about 10- 12 yr old). Mature adults were not caught. Sandbar sharks were captured in all years except 1995 (Fig. 3). The number of individuals collected was highest in 1993. Small juveniles were the domi- nant life stage captured in 1993 and 1997 and larger juveniles in 1994. Young-of-the-year, including neonates, were caught in all years sandbar sharks were captured. The abundance and size of sharks varied with season (Fig. 4). Sandbar sharks were not captured until April when the water temperature approached 22°C. These were mostly small juveniles ranging in size fi-om 800 to 1200 mm TL. Neonates were first captured in June when temperatures reached 25°C and young-of-the-year continued to be caught through October. There was a significant relationship between abundance and water tempera- ture (r-=0.25, P=0.008), but not with salinity or turbidity (P>0.05). Sandbar sharks were most abundant during summer months when all size classes were caught. Few larger juveniles were caught in fall. 30. 25 20- I - 10. i i ji i 0 Young-of-the-Year Q Small juveniles □ Largejuveniles ■ Subadults ^ 500 600 700 800 900 1,000 1.100 1,200 1,300 1.400 1,500 1.600 Total length (mm) Figure 2 Length-frequency distribution of all sandbar sharks (n = 105) by life his- tory stage caught in gill nets and longlines from October 1992 to October 1997. Young-of-the-year includes neonates. Z rv^ [3 Young-of-the-year Q Small juveniles D Largejuveniles ■ Subadults I 1992 1993 1994 1995 Year 1996 1997 Figure 3 Overall abundance of sandbar sharks captured by year and life history stage from October 1992 to October 1997. Young-of-the-year includes neonates. Discussion Presence of neonate and juvenile sandbar sharks in the northeastern Gulf of Mexico suggests that sand- bar sharks pup in the eastern Gulf of Mexico. Springer (1960) proposed the existence of two breeding popula- tions of sandbar sharks, one off the mid-Atlantic coast 390 Fishery Bulletin 97(2), 1999 500 600 700 800 SKXJ 1 ,(X)0 1,1 tX) 1.200 1 .300 1,400 1 .500 1 .600 Summer n=89 Y7/ ^ PT] Young-of-the-year r^ Small juveniles I ) L-argc juveniles ^B Subadults ^ SOO 600 700 800 900 l.OOO .lOO ,200 .300 .•400 .500 I,600 30- 25 - Fall «=7 20- 15- lO- 5- O- ^ F5j r//^ v/y\ r//j 500 600 700 800 S*00 1 .fHK> l.UK) 1 .200 1 , JOG 1 .400 1 .500 1 .1800 mm TL; Sminkey and Musick, 1995) and it is likely that they were able to avoid the sampling gear. Neonate sandbar sharks (24 h. Results Between August 1995 and November 1996, 835 big- eye and 458 yellowfin tuna were released at Cross Seamount. The tagged bigeye tuna were between 40 and 105 cm FL; yellowfin released were between 40 and 90 cm FL. There were no significant differences between the two species in the size distribution of fish tagged and released (Fig. 2) or in size distribu- tion of the recaptures. Because no effort was made to preferentially tag a particular species, the ratio of releases (65'7f bigeye, 2>b'^7c yellowfin) reflected the ratio of species actually caught. Release and re- capture data are summarized in Table 1. The numbers of yellowfin and bigeye tuna recaptured at Cross Seamount were aggregated into 30-day periods of time at liberty and plotted as percentages of to- tal number of fish recaptured. The resultant regression curves for the recapture of each species are shown in Figure 3. An analy- sis of covariance indicated that the slopes of the attrition curves for the two species are signifi- cantly different (P=0.013). The attrition rate (slope of the regres- sion line) for yellowfin tuna is ap- proximately twice the rate for bigeye tuna, and the slopes indi- cate a residence time ( SO'/r recap- tured) of 15 days for yellowfin tuna and 32 days for bigeye tuna. By contrast, the tag attrition Figure 1 Chart of the study area showing position of Cross Seamount in relation to the main Hawaiian Islands. Hxi ■-I . ^~l Fork length (cm) Figure 2 Size distribution of bigeye tuna and yellowfin tuna tagged and released at Cross Seamount. Solid bars = bigeye tuna; open bars = yellowfin tuna. 394 Fishery Bulletin 97(2), 1999 curves for statewide returns (that is, time at liberty for all recaptured Cross Seamount yellowfin and big- eye tuna, regardless of recapture location) were not significantly different (P=0.45, Fig. 3). In addition, there were more recaptures of bigeye tuna with longer periods at liberty than of yellowfin tuna at the seamount and the longest time at liberty was 169 days for bigeye tuna compared with the long- est point-of-release recapture of 93 days for yellow- fin tuna. 100 0 500 10,0- 50- 1.0 05 Q) Q. CO CJ OJ (D o ^_ CD CL 50 100 150 Days at liberty 1000 500 10.0 5.0- 1.0 0,5 A Bigeye Yellowfin 50 100 150 Figure 3 Tag recapture attrition cur\'es (.semi-log plots with 9.5'^/ confidence contours) for bigeye and yellowfin tuna tagged and released at Cross Seamount. (A) All recap- tures from all locations; (B) Recaptured at Cross Seamount. Discussion The main thrust of this analysis was to compare the residence times of the two tuna species at Cross Sea- mount. Consequently, many complicating factors of- ten associated with analysis of tag-and-recapture data could be avoided (especially the impact of vari- ability of fishing effort on the temporal pattern of recaptures) because effort could be assumed to be equal for both species. Therefore, the difference in the tag attrition curves, and the resultant difference in resi- dence times ("half-life"), prob- ably reflect real differences in the behavior of these two spe- cies at Cross Seamount. The similarity of the recapture curves for the entire area (Fig. 3) further suggests that meth- odological or experimental bi- ases are not responsible for the differences in attrition curves obtained for these two species at Cross Seamount. These results differ from those of Fonteneau (1991) who observed no differences in the temporal characteristics of re- captures of tagged bigeye, yel- lowfin, and skipjack tunas re- leased at a seamount in the tropical Atlantic. Fisheries data (Hanamoto, 1976, 1987) and acoustic track- ing (Holland et al., 1990) both indicate that the open ocean behavior of bigeye and yellow- fin tuna is different. Bigeye tuna select colder waters and are therefore usually found deeper than yellowfin tuna, which ori- ent principally to the top of the thermocline and the mixed sur- face layer. However, this verti- cal separation breaks down around fish aggregating devices (FADs, Holland et al., 1990) and floating objects where bigeye tuna move closer to the surface and overlap in vertical distribu- tion with yellowfin tuna. This same effect occurs at Cross Sea- mount where both species are caught in surface schools and bigeye tuna outnumber yellow- 200 B Bigeye — Yellowfin 200 NOTE Holland et a!.: Residence times of Thunnus albacares and T obesus over a seamount 395 fin tuna in the fishery. Similarly, mixed aggi'egations of yellowfin tuna, bigeye tuna, and skipjack tuna have been reported at seamounts in the Atlantic (Fon- teneau, 1991). Although the vertical behavior of yellowfin and bigeye tuna seems to merge at Cross Seamount, the current data indicate that the duration of horizontal orientation to the seamount (as measured by resi- dence time ["half-life"] ) is different. The underlying advantage of seamounts to tuna biology is not well understood. Although seamounts can cause geographically stable regions of planktonic enrichment, it is not known if this enrichment per- sists long enough to move through the trophic chain to the level of the tuna forage base (Boehlert and Genin, 1987). We do know that Cross Seamount is situated in a very dynamic part of the ocean charac- terized by vortices created on the downcurrent side of the main Hawaiian islands (Flament et al.'). As these eddies spin off from the islands, current direc- tion over the seamount can change frequently. If an enriched area of prey does exist, and feeding is the principal underlying reason for tuna aggrega- tions, it is difficult to understand why residence times are different for the two species and quite brief for both. The feeding advantage should impact both spe- cies equally and their residence times at the sea- mount should be similar. A comparison of the stom- ach contents of the two species when caught in sea- mount aggregations would be instructive. It is pos- sible that, rather than acting as feeding stations, seamounts act as orientation points in the larger- scale movement patterns of these fish. Even though they may be too deep for visual detection, seamounts may be recognized by tuna through their ability to detect the effect of seamounts on the earth's mag- netic field (Walker, 1984; Walker et al., 1984; Klimley et al., 1988). The seamounts may act as midocean reference points that may also occasionally harbor increased prey densities, the periodicity and persis- tence of which are driven by events in the surround- ing oceanographic conditions. A navigational role might explain why remote sea- mounts aggregate more tuna than seamounts located closer to land masses (Fonteneau, 1991) and why, in our study, the residence times were quite brief for both yellowfin and bigeye tuna species. The differ- ences in the duration of orientation to the seamount might be explained if the navigational importance of seamounts is different in the broader behavioral rep- ertoires of the two species. ' Flament, P. J., C. Lumpkin, J. Tournadre, P. Kloosterziel, and L.Armi. 1997. Period doubling and vortex pairing in an an- ticyclonic shear flow in the ocean. Manuscript in review. Certainly, the current data indicate that Cross Sea- mount hosts transient populations of both tuna spe- cies rather than long-term populations. This brevity of residence at the seamount for both species probably reduces the chances of excessive fishing exploitation. Acknowledgments We greatly appreciate the expert instruction in tag- ging techniques provided by David Itano, the fish- ing prowess of Tony Frietas, and the role of John Sibert in pointing out the applicability of tag attri- tion curves to these data. This research was funded under Cooperative Agreement No. NA37RJ0199 from NOAA and administered by the Pelagic Fisheries Research Program, Joint Institute for Marine and Atmospheric Research, School of Ocean and Earth Sciences and Technology, University of Hawaii. Literature cited Boehlert, G.W., and A. Genin. 1987. A review of the effects of seamounts on biological processes. In B. H. Keating. P. Fryer, R. Batiza. and G.W. Boehlert (eds.). Seamounts, islands and atolls. Geophys. MonogT. ser. 43, 405 p. Fonteneau, A. 1991. Seamounts and tuna in the tropical Atlantic. Aquat. Living Resour. 4:13-25 Hanamoto, E. 1976. The swimming layer of bigeye tuna. Bull. Jpn. Soc. Fish. Oceanogr. 29:41-44. (Engl. Transl. 21 by T. Otsu, 1977, 7 p. Honolulu Lab. SWFC-NMFS, NOAA.] Hanamoto, E. 1987. Effect of oceanographic environment on bigeye tuna distribution. Bull. Jpn. Soc. Fish. Oceanogr. 51:203-216. Holland, K.N., R. W. Brill, and R. K. C. Chang. 1990. Horizontal and vertical movements of yellowfin and bigeye tuna associated with fish aggregating devices. Fish. Bull. 88:493-507. Kleiber, P., A. W. Argue, and R. E Kearney. 1987. Assessment of Pacific skipjack tuna Katsuwaniis pelamis resources by estimating standing stock and com- ponents of population turnover from tagging data. Can. J. Aquat. Sci. 44:1122-1134. Klimley, A. P., S. B. Butler, D. R. Nelson, and A. T. Stull. 1988. Diel Movements of hammerhead sharks Sphyrna lewini Griffith and Smith, to and from a seamount in the Gulf of California. J. Fish Biol. 33:751-761. Marsac, F., P. Cayre, and F. Conand. 1995. Analysis of small scale movements of yellowfin tuna around fish aggregating devices (FADs) using sonic tagging. Sixth expert consultation on Indian Ocean tu- nas, Colombo, Sri Lanka. 25-29/09/95. 20 p. Walker, M. M. 1984. Learned magnetic field discrimination in yellowfin tuna, Thunnus albacares. J. Comp. Physiol. 155:673-679. Walker, M. M., Kirschvink, J. L, R. K. C.Chang, and A. E. Dizon. 1984. A candidate magnetic sense organ in the yellowfin tuna Thunnus albacares. Science (Wash. D.C.) 224:751-753. 396 Accuracy of at-sea commercial size grading of tiger prawns iPenaeus esculentus and P. semisuicatus) in the Australian northern prawn fishery Michael F. O'Neill David J. Die Brian R. Taylor CSIRO Marine Research Laboratories PO Box 120 Cleveland, Queensland 4163, Australia Present address (for M F ONeill); Southern Fisheries Centre Queensland Department of Primary Industries PQ Box 76, Deception Bay, Queensland 4 508, Australia E-mail address (for M F ONeill) oneillma dpi qld govau) Malcolm J. Faddy University of Queensland Department of Mathematics, Brisbane, Queensland 4072, Australia The size-frequency distribution of the commercial catch is often used as the basis of fisheries stock as- sessments (Pauly and Morgan, 1987; GuUand and Rosenberg, 1992) because most dynamic pro- cesses of populations (growth, sur- vival, recruitment) are reflected in changes in this distribution. The data are generally collected, often at great expense, by sampling the catch at landing sites and markets, or onboard fishing vessels. Size-frequency distributions of prawns iPenaeus esculentus and P. semisuicatus) can also be obtained from fish processors, who grade landings by size. These data are easier and cheaper to obtain than research samples, but unfortunately they are also considered less accurate and lack spatial information. How- ever, they have been used in stock assessment of prawns in Kuwait (Jones and van Zalinge, 1981) and Malaysia (Simpson and Kong, 1978). It is often difficult to relate size data obtained from a processor to time and place of capture of the prawns, but this is not the case when the product is packed onboard, as in Australia's northern prawn fishery (NPF). Trawler operators in the NPF have voluntarily recorded size com- position since 1985, when provision for this was made in operators' daily logbooks (between 30% and 459c of the tiger prawn catch re- ported in the logbooks contain size information). These books are therefore the most comprehensive source of information on the spa- tial and temporal size distribution of the commercial catch of the NPF. Present assessments of the fishery are based on deterministic growth and deterministic seasonal recruit- ment patterns (Wang and Die, 1996) and do not use size-structured data. If available, these data would help relax the assumption of determinis- tic recruitment and improve current stock assessments of the NPF. Before the size data recorded in the logbooks can be used, however, the accuracy of size grading at sea needs to be assessed. This paper examines the accuracy of grading tiger prawns, by using data col- lected from a private firm, A. Raptis and Sons, that operates a large modern processing factory that regularly assesses the onboard grading of product purchased from NPF trawler operators. Although the work presented here relates specifically to the NPF, the practice of onboard size grad- ing is widespread in other fisher- ies around the world. Therefore our methods have potential application to other fisheries. Methods At-sea commercial grading procedures Prawns landed from the NPF are size-graded at sea because both the demand and price structure differ for prawns of different sizes. Com- mercial sizes are based on the num- ber of prawns of the same size per unit of weight (per pound), and the sizes are then grouped in a range to constitute a commercial grade. For example, "9 to 12 grade" means prawns in a range of sizes individu- ally equivalent to between 9 and 12 per pound. The size grades ( especially for the larger sizes) used for tiger prawns are often more precise than those used for other species, and the grades selected by fishermen at sea vary with operator, pack size, and target market. For this project we examined the data for the two pack sizes that were most commonly used during 1993 and 1994: small packs (3 kg) and large, variable weight ( 12-15 kg) packs. Small packs Since the early 1990s, the use of accurate digital scales on many vessels has improved the ac- curacy of procedures for packing prawns into 3-kg or smaller packs, as well as into more conventional larger packs. The sensitivity of Manuscript accepted 12 May 1998. Fish. Bull. 97:396-401 (1999). NOTE O'Neili et al : Accuracy of at-sea commercial size grading of Penaeus spp 397 these digital scales also makes it possible to pack in more precise grades of prawns. Prawns for these small packs are initially sorted by eye by experienced crew, and many are verified by individual weighing. Those prawns that fall out- side a particular size grade are removed and the re- mainder are graded according to corresponding count-per-unit-of-weight tables. Large packs Prawns for onboard grading into the large packs are sorted by eye into groups of about the same size (occasionally by counting the number into a unit of weight [often a pound measured on analogue scales] and grouping them accordingly). Very large and very small prawns are removed and regraded. Quality control assessment in the factory— source of data for analysis A. Raptis and Sons process tiger prawns caught by their own large fleet of trawlers working in the NPF, as well as prawns purchased from other fishermen operating in the same area. The company randomly checks the quality of the packs entering its factory, including the accuracy of the grading. Packs for quality-control assessment were selected at random for every vessel and from all consignments entering the factory. All packs were clearly marked with the vessel's name, the prawn species group, the grade of the prawns, and the date caught. The selected samples were thawed individually. For the small packs, net weight was recorded, and size grading was checked by counting all prawns from each pack and averaging the count. Large and small prawns that did not fit the grade category were se- lected by eye, weighed, and graded individually, and the percentage by weight and the true grade of these prawns were recorded along with the percentage of those correctly graded. With the large packs, a variable 2.5-3 kilogram sample of prawns was randomly taken from each pack, counted, and checked as above. The percent- age by weight and the true grade of incorrectly graded prawns in the sample were recorded. Categorical analysis The results of all factory quality-control checks on both the small and large packs between mid- 1993 and the end of 1994 were examined. Over this time, samples ft-om 51 of the 127 boats that fished different areas in the NPF had been taken. We split the data into three time periods to find out whether the accu- racy of grading early in the year differed from grading Table 1 Commercial size grades (*) used in grading and quality- control assessments of tiger prawns for two different pack sizes. Also shown are the ranges of carapace lengths (mm) for each size grade. Grade (prawn count (per lb) Carapace length (mm) Smal pack Large pack Quality control Under 6 >46 * * * 6 to 8 46-42 * * * Under 10 >39 * 9 to 12 41-36 * * * 10 to 15 38-33 * 13 to 15 35-33 * * * 16 to 20 32-30 * * * 10 to 20 38-30 * Over 20 <29 * 21 to 25 29-28 * 26 to 30 27-26 * 21 to 30 29-26 * that took place later in the year when smaller prawns recruited to the fishery. Period one was fi-om July to December 1993, period two fi-om January to June 1994, and period three ft-om July to December 1994. Data from the large packs, where the grades used were not the same as those for the small packs, were omitted from the analysis because they could not be compared directly. The commonly used size grades for both pack sizes are shown in Table 1, as well as the equivalent carapace length of the prawns. The number of prawns in) contained in the small packs was recovered by converting the net weight of the pack to pounds (weight in kilograms divided by 0.45359), and then by multiplying by the count-per- pound derived from the quality-control inspection. The number of misgraded prawns (r) in the small packs was estimated by multiplying the number of prawns in the sample by the percentage misgraded in that pack. The numbers misgraded for each pe- riod and each size grade were analyzed by fitting binary regression models by means of iterative weighted least squares. The number of misgraded prawns was assumed to have a binomial distribution: Pr n'a-n)" (r=0, l...n) with P( misgraded) = /rand P( correctly graded) = I- n. The probability ir was modelled in terms of the log- odds or logistic transformation (logiit/il - n)). The computed deviance statistic, approximately distrib- uted as chi-squared, was used in goodness-of-fit tests. 398 Fishery Bulletin 97(2), 1999 It was not possible to recover the total number of prawns in the large packs because the variable sample weight (between 2.5 and 3 kilograms) was not recorded in the Raptis database. Therefore esti- mates of the number of misgraded prawns were de- rived only for the samples and not for the whole pack. Because these samples were randomly chosen, it was possible to assume that the assessment of grading accuracy was representative of the grading accuracy for the whole pack. For the large pack samples, the proportion mis- graded had a mean (Eq. 1) and variance (Eq. 2) over the different samples Mean = n and Variance = E ;rfl-;r) (1) (2) where n = the misgrading probability; and n = the sample size. Because the weight range of the samples was small, it was possible to estimate the expected reciprocal sample size n (Eq. 4) by integrating over the sample weight range (assumed for mathematical convience to be uniformly distributed between 2.5 kg and 3 kg) (Eq. 3): E n j J2.5 3 0.45359 2dw piv f0.45359x2xln(3/2.5)'| (3) (4) Results Small packs Of the 21,443 tiger prawns in 293 small packs that were assessed, an estimated 1937 (9%) prawns in 229 packs were misgraded. There were significant changes in the proportion misgraded with both pe- riod of catch and size grade, with higher proportions of misgraded prawns in the small size grades (Table 2; Fig. lA). Overall, grading accuracy tended to in- crease over the 18 months examined (Fig. lA). The size of the misgraded prawns over the differ- ent grades did not show a consistent pattern, but generally larger prawn grade packs tended to con- tain smaller prawns (Fig. lA). The proportion of misgraded prawns that should have been in smaller grades, however, was 1 not constant over all size grades within each pe- riod of catch (Table 3; Fig. lA); 2 not the same for each size grade over the three periods examined (Table 4; Fig. lA). Of the misgraded prawns, 99^^ were size-graded either one grade larger or one grade smaller. Only grades 9 to 12 and 16 to 20 contained prawns misgraded by as much as two size grades, with no more than 2'/( so misgraded. Because there was no larger grade, prawns misgraded in the under 6 size, were graded as 6 to 8. If the proportion of prawns graded size i by fisher- men at-sea that were actually sizejii, j=l for under 6, 2 for 6 to 8, 3 for 9 to 12, 4 for 13 to 15, 5 for 16 to 20, and 6 for over 20 prawns per pound) obtained from the sample data are denoted by 9^ , then the proportions, p^, of all prawns graded as size ; at-sea can be adjusted with the equation: f^PA, (5) where p = count per pound; 0.45359 kg = 1 lb; and iv = sample weight. The relationship between Equations 1 and 2 here is the same as that for the binomial distribution; there- fore the data were analyzed by fitting binomial re- gression models. The size of misgraded prawns was examined to determine whether misgrading was a result of in- cluding small prawns in large grades or vice versa. The number of size grades in which misgrading oc- curred was also assessed. to give a corrected grade size distribution (j=l, 2, ..., 6). Shown in Table 5 are the corrected distributions compared with at-sea grading for the small packs. The adjustments can be seen to be quite modest, and the at-sea gradings provide a reliable assessment of the size distribution. Large packs Samples containing an estimated 8210 tiger prawns from 124 large packs were assessed. Of these samples, an estimated 2914 (35%) prawns from 107 packs were misgraded. Again, there were significant NOTE O'Neill et a\. Accuracy of at-sea commercial size grading of Penaeus spp. 399 0.6 0.5 0.4 -I 0.3 0.2 0.1 0 Period 1 : July - December 1993 U/6 6/8 9/12 13/15 16/20 0.6 Period 2: January - June 1994 -a 0.5 -n a Of) 0.4 ly. p c 0.3 - o o 0.2 & f; a. 0.1 oJ u/6 6/8 9/12 13/15 16/20 Period 3; July - December 1994 0.5 0.4 0.3 0.2 0.1 ^ 0 U/6 6/8 9/12 13/15 16/20 0.6 0.5 0.4 0.3 0.2 0.1 0 Size grade 9/12 13/15 16/20 - T - ^ ■ u/6 6/8 9/12 13/15 16/20 U/6 6/8 9/12 13/15 16/20 Figure 1 The proportion ( +SE ) of misgraded tiger prawns from each size grade, for (A) the small packs and (B) the large packs. The proportion misgraded was split between prawns that were misgraded too large (white) and those that were misgraded too small (gray). changes in the proportion misgraded with period of catch and size grade, with significant differences in the proportion misgraded with period of catch, and generally higher proportions of the smaller size grades misgraded (Table 2; Fig. IB). There was a tendency for smaller-prawn grade packs to contain larger prawns (Fig. IB). The pro- portion of misgraded prawns that should have been in a larger-size grade, however, was 1 not constant over all size grades, within any one period of catch (Table 3; Fig. IB); 2 not the same for each size grade, over the three periods examined (Table 4; Fig. IB). 400 Fishery Bulletin 97(2), 1999 Table 2 Binomial model fits for each period of catch (1: July-December 1993, 2: January-June 1994, and 3: July-December 1994) and pack-size combination, (a, b, c) denote groups of size grades with no significant differences in the proportion of misgraded prawns. Size grade Pack Time size period Under 6 6 to 8 9 to 12 13 to 15 16 to 20 x^ df P Small a a a 1 b b 1.517 3 0.68 a a 2 b 0.476 1 0.52 c c a 3 b b 3.244 2 0.20 c c Large a 1 b - ' c a a 2 b b b 2.695 3 0.44 a a 3 b 2.892 2 0.24 c c ' Zero degrees of freedom. Table 3 Chi-squared statistics for the constant model of misgraded prawn size in each period of catch 1 1: July-December 1993, 2: January-June 1994, and 3: July-December 1994), for both small and large packs. P< 0.0001 Period Small pack Large pack 45.088 45.916 32.915 17.963 55.701 83.060 Table 4 Chi-squared statistics for the constant model of misgraded prawn size in each size grade for both small and large packs. *P< 0.0001, ■P= 0.13, P= 0.03. Size grade Small pack Large pack 6 to 8 4.1494 ' 14.408 9 to 12 33.271 4.7639 ■ 13 to 15 220.58 ■ 27.039 16 to 20 16.159 106.83 Of the prawns misgraded, 92'^J were sized either one grade larger or smaller. Of those misgraded in the 16 to 20 grade, 11% were misgraded by as much as two grades, whereas less than 2% were so misgraded for the other size grades. The misgrading proportions can again be used in Equation 5 to obtain the at-sea grade-size distribu- tion for these large packs. Shown in Table 5 are these corrected distributions compared with at-sea gradings; here the adjustments can be seen to be more substantial than those for the small packs, par- ticularly for the smaller grades 13 to 15 and 16 to 20 prawns per pound, owing to the tendency of the fish- ermen to classify the prawns to smaller size grades. Discussion Our analysis indicates that small prawns are graded less accurately than large ones. Given that the length NOTE O'Neill et al : Accuracy of at-sea commercial size grading of Penaeus spp. 401 Table 5 Size grade distributions, expressed as proportions , from grading at sea and adjusted according to Equation 5 Period 1 Period 2 Period 3 Size grade (July-December 1993) (January-June 1994) (July-December 1994) at-sea adjusted at-sea adjusted at-sea adjusted Small packs Under 6 0.15 0.14 0.012 0.015 0.098 0.097 6 to 8 0.16 0.17 0.19 0.19 0.11 0.11 9 to 12 0.30 0.33 0.31 0.30 0.32 0.30 13 to 15 0.26 0.23 0.28 0.28 0.24 0.23 16 to 20 0.13 0.11 0.21 0.20 0.24 0.24 Over 20 0 0.015 0 0.014 0 0.018 Large variable packs Under 6 — — 0.11 0.071 0.12 0.15 6 to 8 — — 0.059 0.11 0.17 0.15 9 to 12 0.079 0.12 0.12 0.16 0.22 0.28 13 to 15 0.055 0.43 0.10 0.22 0.14 0.21 16 to 20 0.87 0.42 0.61 0.38 0.36 0.20 Over 20 0 0.030 0 0.069 0 0.017 range corresponding to the small commercial grades is narrow (Table 1 ), it is perhaps not surprising that small prawns tend to be misgraded more frequently. Alternatively the grading of small prawns may be less accurate because they are less valuable than large prawns and therefore less time is spent on grad- ing each individual. In small packs, misgraded prawns were generally gi'aded into larger categories, whereas in large packs, those misgraded were generally placed into smaller categories. Incorrectly graded prawns from both pack sizes, however, tend to be incorrectly graded by only one size category, so that all prawns were graded to within three and six millimeters carapace length of their corresponding size grade. The high proportion of landings graded and the accuracy of some of this gi'ading suggest that size information contained in the NPF logbooks could be valuable for stock assessment. Most prawns sold in small (3-kg) packs have been accurately graded by fishermen, and these gradings could be used as a reasonable measure of the length-frequency distri- bution of the prawns. However, prawns sold in larger (12-15 kg) packs are graded less accurately, espe- cially for the smaller grade sizes, and it is recom- mended that data from quality inspections be used to correct fishermen's grade-size distribution. Although the work outlined in this paper was done for the Australian northern prawn fishery, similar analyses using similar methods could be carried out for other fisheries if comparable data on size gradings were available. Acknowledgments We thank A. Raptis and Sons and, in particular, David Crighton and Paul Hand for providing the quality-control data set and describing the quality- control procedures. Ted Wassenberg, Chris Jackson, and Carolyn Robins made constructive comments on an earlier version of this manuscript. Literature cited GuUand, J. A., and A. A. Rosenberg. 1 992. A review of length-based approaches to assessing fish stocks. FAO Fisheries Tech. Paper 323, FAO, Rome, 100 p. Jones, R., and N. P. van Zalinge. 1981. Estimates of mortality rate and population size for shrimp in Kuwait waters. Kuwait Bull. Mar Sci. 2:273-88. Pauly, D., and G. R. Morgan. 1987. Length-based methods in fisheries research. ICLARM Conference Proceedings 13. 468 p. Simpson, A. C, and C. P. Kong. 1978. The prawn fisheries of Sabah, Malaysia. Ministry of Agriculture, Malaysia. Fish. Bull. (Malaysia) 22, 25 p. Wang, Y-G., and D. J. Die. 1996. Stock-recruitment relationships of the tiger prawns iPenaeus esciilentus and Penaeus semisulcatus) in the Aus- tralian Northern Prawn fishery. Mar. Freshwater Res. 47:87-95. 402 Digestive tract parasites in rhynchoteuthion squid paralarvae, particularly in ///ex argentinus (Cephalopoda: Ommastrephidae) Erica A. G. Vidal Manuel Haimovici Departamento de Oceanografia Universidade do Rio Grande (FURG) C- P, 474 Rio Grande RS 96 201-900, Brazil E-mail address (for E A G Vidal) eavidalauimbedu Present address (for E. A G. Vidal): National Resource Center for Cephalopods Marine Biomedical Institute University of Texas Medical Branch 301 University Blvd., Galveston, Texas 77555-1163 Cephalopods play an important role in the transfer of parasites through the food web. They are second and third intermediate hosts for larval stages of digeneans, cestodes, and nematodes. The final hosts are usu- ally fishes, sea birds, and marine mammals (Hochberg, 1990). Nev- ertheless, the interactions of cepha- lopods and their parasites are still poorly understood, and almost nothing is know about parasitism in cephalopod paralarvae and early juveniles (Vecchione, 1987). Infor- mation on parasitism in the early stages of cephalopod life cycles can provide a better understanding of host-parasite relationships as well as the environment and conditions in which infection first occurs. Ommastrephid paralarvae are among the smallest cephalopods at hatching and are termed "rhyncho- teuthion" because their tentacles are fused into a proboscis-like structure. Three distinct morphological types of rhynchoteuthion paralarvae (types "A," "B," and "C") are found commonly off southern Brazil (Haimovici et al., 1995). "Type C" is Illex argentinus (Castellanos, 1960) and "type A" probably is Ommastrephes bartramii (Lesueur, 1821). Illex ctrgentinuf: is distrib- uted in the southwestern Atlantic Ocean from 22" to 54°S (Roper et al., 1984), and it currently supports the largest squid fishery in the world. This species is an important link in the trophic relations of the pelagic ecosystem of this region and constitutes a major element of the diet of several commercial fishes, such as Thunnus obesus, Merluccius hi/bbsi, Xiphias gladiiis, and Poly- prion americanus (Santos, 1992; Ivanovic and Brunetti, 1994). The digestive tract contents of Illex argentinus and the other rhyn- choteuthion types were examined by Vidal and Haimovici (1998). During examination of the diges- tive tracts, observations on parasit- ism were made. In this note we re- port the parasites (protistans and metazoans) found in the digestive tract of rhynchoteuthion paralarvae, mainly Illex argentinus. Material and methods Ommastrephid paralarvae were col- lected in four surveys carried out be- tween Cape of Santa Marta Grande (28"30'S)and Chui (34 20'S), Brazil, during the spring of 1987, summer of 1990, autumn of 1991. and win- ter of 1988. Paired bongo nets, with 60-cm mouth diameter and 0.33- mm mesh were used in oblique tows between the surface and depths of approximately 300 m. This gear was towed at a speed of 2 kn and deployed for 5-15 min. Afterwards, paralarvae were fixed and pre- served in formalin. Digestive tracts were examined from paralarvae of 72 I. argentinus ( 1.0-8.0 mm of mantle length (ML)), 12 "type A" ( 1.0-5.8 mm ML) and 4 "type B" (2.0-4.0 mm ML). Para- larvae were stained with alcian blue and then cleared with trypsin, following the method of Vecchione (1991). This method makes the paralarvae semitransparent and the parasites in the gut easier to see. The stomach, caecum, and in- testine of all paralarvae were ex- amined for parasites with a light microscope at a magnification of 400x. The esophagus was not exam- ined. Five /. argentinus juveniles (16.0-38.0 mm ML) and 14 para- larvae (1.6-9.7 mm ML) of/, argen- tinus with no previous clearing and staining treatment were also exam- ined. The prevalence and the mean intensity of infestation (.number of a particular parasite per infected individuals) were calculated ac- cording to Margolis et al. ( 1982). Results Parasites were not found in the 14 untreated paralarvae. The opaque walls of the internal organs made the visualization of parasites diffi- cult. A single copepod parasite was found in one of the five untreated juveniles. Four types of parasites were ob- sei-ved in paralarvae and juvenile /. argentinus: coccidians (Apicom- plexa; Sporozoea: Eucoccidiida), didymozoid-metacercariae ( Platy- helminthes: Trematoda: Digenea), Manuscript accepted 19 May 1998. Fi.sli. Bull. 97:402-405 (1999). NOTE Vidal and Halmovici: Parasites in rhynchoteuthion squid paralarvae 403 Table 1 Results of examinations of rhynchoteuthion paralarvae from four survevs conducted between the spring of 1987 and autumn of 1991. + = presence; 0 = absence. Summer Spring 1987 1990 Autumn L991 Winter 1988 Illex argentinus "Type B" "Type A" Illex argentinus "Type B" Illex argentinus Paralarvae mantle length (mm) 1.0-8.0 2.8 1.0-5.8 1.3-5.5 2.0-4.0 1.0-7.8 Digestive tracts examined 41 1 12 19 3 12 Parasites: Nematode + 0 0 0 0 0 Aggregata sp. (coccidiani 0 0 0 0 0 + Didymozoids Moniticaecum sp. + 0 + + 0 + Prevalence (%) 7.3 72.0 10.5 8.3 Mean intensity of infestation 2. .3 :i ,s 1 .i 10 a nematode (Nematoda) and a copepod (Arthropoda: Crustacea: Siphonostomatoida). About 50 coccidians (Aggregata sp. ) were found distributed along the cae- cum wall of a 7.8-mm-ML paralarva collected in win- ter. Because the paralarva was preserved in glycerin for several months, taxonomic details of the para- sites were not very clear. Cyst capsules (sporocystes) varied in size from 11 to 17 |.im. However, the de- tailed structures inside the capsules were obscure and no certainty could be attained about the pres- ence of sporozooites; hence this protozoan was as- signed to the genus Aggregata. The observed para- sites may have been in a very early stage of develop- ment and if the capsules actually represent sporo- cystes they were unusual because only 1 or 2 were present inside an amorphous oocyst mass (Hoch- bergM. These parasites are deposited at the Santa Barbara Museum of Natural History. Six/, argentinus paralarvae (2.2-6.7 mm ML) from autumn, winter, and spring surveys contained 1-3 didymozoid metacercariae, probably type Monili- caecum. The body dimensions of the parasites ranged from 140 to 200 |am and because of their size and the absence of testicular anlagen, they appeared to be recently hatched. A single trematode attached to the caecum wall was found in the smallest paralarva. Three parasites were found in the largest paralarva, two attached to the stomach wall and one near the buccal mass. In the other paralarvae, the parasites were found attached to the caecum wall. Data on prevalence and mean intensity of infestation of dif- ferent paralarval types and surveys of didymozoids ' Hochberg. F. G. 1997. Santa Barbara Museum of Natural History, 2559 Puesta del Sol Road, Santa Barbara, CA 93105. Personal commun. are presented in Table 1. We found a single (nematode attached to the caecum wall of a 7.0-mm-ML rhyncho- teuthion from the spring sample. The copepod para- site was located on the caecum wall of a 38.0-mm-ML juvenile obtained in spring. The copepod was identi- fied as an adult male of Metacaligus uruguayensis ( Siphonostomatoida: Caligidae ) ( Ho and Bashirullah, 1977). Illex argentinus is a newly recorded host for M. uruguayensis. In the summer of 1990, didymozoid metacercariae were found in "type A" rhynchoteuthions from 2.0 to 5.0 mm ML. Prevalence was 729f and the mean in- tensity of infestation was 3.5 (Table 1). All were at- tached to the caecum or stomach walls, except for three parasites attached to the inner wall of the mantle and two others to the caecum wall of a single paralarva (3.6 mm ML). Parasites were not found in "type B" rhyncho- teuthions. However, only four paralarvae of this type were examined. Discussion Illex argentinus first acquire didymozoid metacer- cariae as small rhynchoteuthions. Hatchlings typi- cally are found off southern Brazil at the outer shelf and slope in tropical and subtropical waters (Vidal and Haimovici, 1997). The infestation may be related to the fact that paralarvae are distributed mainly in tropical waters, where cystophorous cercariae may be abundant and not necessarily because the paralarvae are eating infected intermediate hosts. Crustacean prey were not found in the digestive tracts of rhynchoteuthions smaller than 3.7 mm ML (Vidal and Haimovici, 1998). However, paralarvae 404 Fishery Bulletin 97(2), 1999 larger than this size could also be infested by con- suming infected second-intermediate hosts, such as copepods and other crustaceans. Infestation in the smaller rhynchoteuthions may occur when free cystophorous cercariae are released into the plankton and passively enter the mantle cav- ity during respiration ( Gaevskaya, 1976 ) or when the cercariae attach to mucus that covers the mantle and head of rhynchoteuthions. They may also be ingested directly, in the same way that crustaceans are in- fested. If this is true, ommastrephid paralarvae may act as second intermediate hosts for some didy- mozoids. The presence of didymozoid metacercariae in the digestive tracts of rhynchoteuthions as small as 2.0 mm ML shows that the infection occurs well before that previously recorded by Gaevskaya and NigmatuUin (1983). Additionally, the size of the didymozoid metacercariae type Monilicaecum (140- 200 |.im) found on rhynchoteuthion paralarvae is strong evidence that they are in an early develop- mental stage in relation to those found on adult /. argentinus from southern Brazil (400-800 |.im) (Santos, 1992). Our results are in accordance with the suggestion of Hochberg ( 1990) that didymozoids can grow inside the squid host. Hochberg ( 1990 ) also reported that a peak of infestation by didymozoids occurs in squids from 10 to 25 cm in ML. Nevertheless, infection decreases in adult individuals ( Gaevskaya and NigmatuUin, 1977. 1983) owing to the fact that the didymozoids die after they reach a maximum size in a specific host (Hochberg, 1990). Larger squids may be reinfected by didymozoid metacercariae when they feed on infested fish prey (Hochberg, 1990). Our re- sults indicate that more paralarvae should be exam- ined in the future. If it is correct to assume that re- cently hatched didymozoid metacercariae usually infest early paralarvae, information on size and de- velopmental stage of both host and parasites would help to understand their relationship better. Didymozoid metacercariae were found in /. argen- tinus rhynchoteuthions from autumn to spring and in "type A" rhynchoteuthions during the summer. This morphotype had a higher prevalence and infes- tation of didymozoids than those of/, argentinus. They were collected in the summer of 1990, when tropical waters dominate the study area (Vidal and Haimovici, 1997). Thus, these larval trematodes are present off southern Brazil throughout the year This finding is in agreement with the occurrence of the high prevalence and intensities of didymozoid meta- cercariae in juvenile and immature /. argentinus from the same area year around (Santos, 1992). Only the sexual stages ofAggregata occur in cepha- lopods, whereas asexual stages infect the digestive tracts of crustaceans (Hochberg, 1990). Aggregata appears to be host-specific only in cephalopods, not in crustaceans, and is a common parasite of Sepia and Octopus (Hochberg, 1990). Our study is the first to report its presence in /. argentinus. This coccidian has not been found in adults of/, argentinus collected off southern Brazil (Santos, 1992) or Argentina (NigmatuUin and Shukhgalter, 1990; Sardella et al., 1990 ). Martial ia hyadesi, a southern Atlantic oceanic species, and Todaropsis eblanae and Todarodes sagit- tatus in waters off Spain are the other ommastrephid squids in which Aggregata have been reported (Gaevskaya et al. 1986, Pascual et al. 1996). It is important to stress that infections hy Aggregata can only be established after paralarvae begin to feed on crustaceans infected with the asexual stages of the parasite. The infected paralarva observed in this study was 7.8 mm ML; however, crustacean prey, mainly copepods, have been found in the gut of /. argentinus paralai-vae larger than 3.7 mm ML (Vidal and Haimovici, 1998). Metacaligus uruguayensis was originally described from the gill cavity of Trichiurus lepturus from Ven- ezuela (Ho and Bashirullah, 19771. This genus pre- viously has not been reported from cephalopods (see review by Hochberg, 1990). Associations between copepods and cephalopods appear to be primarily commensal and not truly parasitic (Hochberg, 1990). In southern Brazil, small and large adult T. lepturus (70-1000 mm TL) occasionally prey on juvenile and adult/, argentinus (22-200 mm ML) (Martins, 1992); however, additional information is required to de- termine whether /. argentinus serves as a final or only an accidental host for this copepod. Acknowledgments We would like to thank F. G. Hochberg for his valu- able review of the manuscript and for examining the coccidian parasites. We also thank G. A. Boxshall for his support and for the identification of the copepod parasite. E. A. G. Vidal was supported by the Brazil- ian National Research Council (CNPq). Literature cited Castellanos, Z. J. A. 1960. Un:i nueva especie de calamar Argentino Omnia- strcphcs argentinus sp. nov. (Mollusca. Cephalopoda). Nt'olropica 6( 20):.5.5-58. (laevskaya, A.V. 1976. On tht> hflminthofauna of the Atlantic squid Omma- stifplii's bartramii Le.sueur. In Biological fisheries re- search in Atlantic Ocean. AtanNIRO Works 69:89-96. |In Russian.] NOTE Vidal and Halmovici; Parasites in rhynchoteuthion squid paralarvae 405 Gaevskaya, A.V., and Ch. M. Nigmatullin. 1977. Distribution of the metacercariae of didymozoid trematodes among Atlantic squids of the family Omma- strephidae. In All-Union scientific conference on the utili- zation of commercial invertebrates for food, fodder and tech- nological purposes, Odessa, p 20-22. [Abstract, in Russian.] 1983. Ecological aspects of age variability of the helmin- thofauna of squids of the family Ommastrephidae. /?! Conference on the biological bases of the control of helm- inths of animals and plants. Moscow, p 19-21. [Abstract.] Gaevskaya, A. V., Ch. M. Nigmatullin, and O. A. Shukhgalter. 1986. Comparative ecological characteristics of the para- sitofauna of common species of squid of the family Omma- strephidae in the southwestern Atlantic. In Abstracts of the proceedings of the 4th All-Union conference on com- mercial invertebrates. Sevastopol, p 337-338. [In Russian.] Haimovici, M., E. A. G. Vidal, and J. A. A. Perez. 1995. Larvae of I Ilex argentmus from five surveys on the continental shelf of southern Brazil. ICES Mar. Sci. Symp. 199:414-424. Ho, J. S., and A. K. M. Bashirullah. 1977. Two species of caligid copepods (Crustacea) parasitic on marine fishes of Venezuela, with a discussion of Metacaligus Thomsen, 1949. J. Nat. Hist. 11:703-714. Hochberg, F. G. 1990. Diseases caused by protistans and metazoans. In O. Kinne (ed.). Diseases of marine animals. Ill: Introduc- tion, Cephalopoda, Annelida, Crustacea, Chaetognatha, Echinodermata, Urochordata, p. 47-227. Biologische Anstalt Helgoland, Hamburg. Germany. Ivanovic, M., and N. E. Brunetti. 1994. Food and feeding of Illex argentinus. In P. G. Rodhouse. U. Piatkowski, and C. C. Lu, (eds.) Southern Ocean cephalopods: life cycles and populations. Antarctic Science 6: 18.'j-193. Margolis, L., G. W. Esch, J. C. Holmes, A. M. Kuris, and G. A. Shad. 1982. The use of ecological terms in Parasitology. J. Parasitol. 68:131-133. Martins, A. S. 1992. Bioecologia do peixe espada Tnchiurus lepturus. Lin- neaus, 1758. no sul do Brasil. Tese de Mestrado, Universi- dade do Rio Grande (FURGi, Rio Grande RS, Brasil, 149 p. Nigmatullin, Ch. M., and O. A. Shukgalter. 1990. Helmintofauna y aspectos ecologicos de las relaciones parasitarias del calamar {Illex argentinus) en el Atlantico Sudoccidental. Frente Maritimo 7:57-68. Pascual, S., C. Gestal, J. M. Estevez, H. Rodriguez, M. Soto, E. Abollo and C. Arias. 1996. Parasites in commercially-e.xploited cephalopods (MoUusca, Cephalopoda! in Spain: an updated perspective. Aquaculture 142:1-10. Roper, C. F. E., M. Sweeney, and C. E. Nauen. 1984. FAO species catalogue. Vol. 3. Cephalopods of the world. FAO Fish. Synop. 125. FAO, Rome, 227 p. Santos, R. A. 1992. Relagoes trbficas do calamar argentino Illex argen- tinus (Castellanos, 1960) (Teuthoidea: Ommastrephidae) no sul do Brasil. Tese de Mestrado. Universidade do Rio Grande (FURG). Rio Grande-RS. Brasil. 83 p. Sardella, N. H., M. I. Roldan, and D. Tanzola. 1990. Helmintos parasitos del calamar {Illex argentinus) en la subpoblacion bonaerense-norpatagonica. Frente Maritimo 7: 53-56. Vecchione, M. 1987. Juvenile ecology. In P. R. Boyle (ed.), Cephalopod life cycles. II, p. 61-84. Academic Press, London. 1991. A method for examining the structure and contents of the digestive tract in paralarval squids. Bull. Mar. Sci. 49:300-308. Vidal, E. A. G., and M. Haimovici. 1997. Distribution and transport of Illex argentinus para- larvae across the western boundary of the Brazil-Malvinas confluence front off southern Brazil. In 63rd annual meet- ing— America Malacological Union and 30th annual meet- ing— Western Society of Malacologists, Santa Barbara, California, p 61. [Abstract.] 1998. Feeding and the possible role of the proboscis and mucus cover in the ingestion of microorganisms by rhyncho- teuthion paralarvae (Cephalopoda: Ommastrephidae). Bull. Mar. Sci. 63(2):305-316. 406 Diet of Pacific sleeper shark, Somniosus pacificus, In the Gulf of Alaska Mei-Sun Yang Benjamin N. Page Alaska Fisheries Science Center 7600 Sand Point Way NE Seattle, Washington 98115-0070 E-mail addre5s Mei-Sun Yangffnoaa gov The sleeper shark, Somniosus paci- ficus, ranges from Chile (Crovetto et al., 1992 ) through southern Cali- fornia (Phillips, 1953), British Columbia to the Gulf of Alaska (Bright, 1959), the Bering Sea (Wilimovsky,1954), and Japan (Tanaka et al.,1982). It is thought to be a voracious and versatile feeder and its diet has been shown to include young marine mammals, such as harbor seal, Phoca vitulina (Bright, 1959), and the southern right whale dolphin, Lissodelphis peronii, (Crovetto et al.,1992). The purpose of this study is to describe the diet of the sleeper shark in the Gulf of Alaska area. Methods Sleeper sharks were collected in the Gulf of Alaska (Fig. 1) between June and August 1996 from the longline vessel A/as/?a Leader, and the bottom-trawl vessels Vester- aalen and American No. 1. After dissections onboard, each indi- vidual stomach was put in a plas- tic bag and frozen at sea. Tags re- cording date, location, and length and sex of shark were included in the sample bag. Information on the station, shark total length (TL) (measurements follow Castro, 1983), and dates samples were col- lected are listed in Table 1. Stom- achs were thawed in the laboratory and stomach contents were ana- lyzed. Prey were identified to the lowest taxonomic level possible. Each prey item was weighed and standard length of prey fish was measured. Percent frequency of oc- currence and the percentage of the total weight of each prey item were calculated. Octopus beak measure- ments were made to estimate live wet weights of octopi. According to Robinson and Hartwick ( 1983), pig- ment-upper-lateral-wall length (PULWL) has the best correlation coefficient with the live wet weight of the North Pacific giant octopus. Octopus dofleini; we therefore used these PULWLs to estimate the live wet weight of the octopus. Live wet weights of Octopus dofleini that had been consumed were calculated by using Robinson and Hartwick's (1983) equation: In (PULWUmm)) = 0.274 In {weight(\ig) -(-2.674. (1) Results A total of 13 sleeper shark stomachs were analyzed; two were empty, 11 contained food (Tables 1 and 2). The length of the sleeper sharks ranged from 218 cm to 295 cm TL (mean= 264.5 cm; SD=24.9 cm. Arrowtooth flounder, Atheresthes stomias, was the most important prey, represent- ing &!'/( of the total stomach con- tent weight (64*^ of frequency of occurrence). The size of arrowtooth flounder consumed by sleeper sharks ranged from 38 cm to 65 cm TL, (mean=44.8 cm; SD=8.0 cm). Other prey included a 48-cm wall- eye pollock, Theragra chalcogram- ma (5.2% by weight), a single 33- cm rockfish, Sebastes sp., a 40-cm Pacific salmon, Oncorhynchus sp., and a 26-cm flathead sole, Hippo- glossoides elassodon, as well as three unidentifiable flatfish. Octo- pus dofleini were the most impor- tant invertebrate found in the diet of sleeper sharks, representing 5% of the total stomach contents weight and 73% of the frequency of occurrence. The estimated wet weight o^ Octopus dofleini (based on beak measurements) ranged from 5.65 kg to 29.07 kg (mean=18.51 kg; SD=6.58 kg). The PULWL mea- surements ranged from 23.3 mm to 36.5 mm (mean=31.78 mm; SD= 3.64 mm). Less important inverte- brate prey included squids, snails (Fusitriton sp.), hermit crabs, and gammarid amphipods. Fish offal (five arrowtooth flounder heads) was found in one sleeper shark stomach. It represented 12% of the total stomach contents weight and had a frequency of occurrence of 9% . No Steller sea lion parts were found in the 13 sleeper shark stom- achs examined. In our study, three specimens were collected from bottom trawls, ten from longline surveys. The weight of stomach contents of the three specimens collected from bot- tom trawls (sharks no. 4, 12, and 13) were more than 2000 g (4506 g; 2321 g; and 11,782 g, respectively), whereas weight of stomach con- tents of the specimens collected from long lines were much lower than those collected from the bot- tom trawls (only three weighed more than 500 g, one weighed more than 3000 g, and the rest weighed less than 50 g). Less food in the stomachs of the sleeper sharks col- Manuscript accepted 13 May 1998. Fi.sh. Bull. 97:406-409 (1999). NOTE Yang and Page: Diet of Somniosus pacificus 407 Figure 1 Locations of sleeper sharks collected in Gulf of Alaska Table 1 Some information on the 13 sleeper shark. Somniosus pa ClflCUS stomachs collected in the Gulf of Alaska in 1996. TL = Total length; Wt = stomach contents weight. Date Vessel Station no. Depth (m) Shark no. TL (cm) Wt(g) 24 Jun '96 Alaska Leader 149 241 7 274 520 24 Jun '96 Alaska Leader 249 253 11 218 12 25 Jun '96 Alaska Leader 250 267 2 292 39 25 Jun '96 Alaska Leader 250 267 5 287 3094 25 Jun '96 Alaska Leader 250 267 6 249 0 25 Jun '96 Alaska Leader 150 240 8 284 821 26 Jun '96 Alaska Leader 151 240 1 295 0 26 Jun '96 Alaska Leader 151 240 3 274 11 26 Jun '96 Alaska Leader 251 255 9 244 11 27 Jun '96 Alaska Leader 248 263 10 244 900 28 Jun '96 Vesteraalen 155 86 4 274 4506 3 Aug '96 American No. 1 1 101 12 229 2321 4 Aug '96 American No. 1 2 113 13 274 11.782 lected by longlines is probably caused by regurgita- tion during the long operation (two to six hours) of longline surveys. Our data also indicate that more prey items (e.g. walleye pollock, salmon, rockfish, and snails) were found in stomachs collected by bottom trawls than collected from longlines (mainly arrow- tooth flounder and octopus). Bottom depths (Table 1) for the samples collected from bottom trawls were shallower (from 86 to 113 m) than for samples col- lected from longlines (from 240 to 267 ml; bottom depth may also be a factor in the diet variations. Length of the sleeper sharks (Table 1) might also affect their diet; owing to the small sample size, how- ever, no comparisons could be made from our study. 408 Fishery Bulletin 97(2), 1999 Discussion On the basis of our sampling of stomach contents, sleeper sharks appear to feed mainly on the bottom. Even though Octopus dofleini represented only 5% of the total stomach content weight, they occurred in a high percentage of sleeper shark stomachs ( 739^ ). Other researchers have also found that benthic fish and invertebrates are the predominate species in the diet of sleeper sharks. Phillips ( 1953) found a sleeper shark in California that had fed on rockfish. Gotshall and Jow (1965) found that the diet of sleeper shark included rex sole, Glyptocephalus zachirus; Dover sole, Microsto77Jus pacificus; Pacific halibut, Hippoglossus stenolepis: and cephalopods. The diet of sleeper sharks vary with their size. Gotshall and Jow (1965) described the main food of a 114-cm female sleeper shark as Moroteuthis ro- busta. In our study lengths of sleeper sharks ranged between 200 and 300 cm. Their diets consisted mainly of arrowtooth flounder, walleye pollock, and cepha- lopods. Larger sleeper sharks (360-400 cm) have been reported to consume not only fishes and cepha- lopods, but also marine mammals, i.e. harbor seal (Bright, 1959) and southern right whale (Crovetto etal., 1992). We would like to note that stomach samples used in our study were from the area southwest of Kodiak Island, close to Steller sea lion iEumetopias jubatus) rookeries at Chowiet and Chirikof Islands, as well as near numerous sea lion haulouts (Sease et al., 1993; NMFS, 1995). However, sleeper shark attacks on pups have not been reported at any time of year and we did not find evidence of predation on sea li- ons in our study. We did find fish offal (five arrowtooth flounder heads) in stomach samples. Bigelow and Schroeder ( 1948) also reported that Greenland shark, Somniosus microcephalus (a similar congeneric spe- cies of sleeper shark in the Atlantic Ocean), devours carrion, such as whale meat and blubber from whal- ing operations. It seems that sleeper shark, like Greenland shark, is sluggish and likes to stay on the bottom, feeding opportunistically on what they en- counter in the environment (including carrion and fish offal). Acknowledgments We would like to thank Lowell Fritz, Patricia Livings- ton, John Sease, and three anonymous referees for reviewing this manuscript. Table 2 Prey items (expressed in percent frequency of occurrence (7fF0), and percent total weight CXW)) of sleeper shark, Somniosus pacificus, collected in Gulf of Alaska in 1996. Prey name %F0 %W Gastropod (snail) 9.09 0.49 Fusitriton sp. i snail) 9.09 0.19 Cephalopod (squid and octopus) 27.27 0.17 Teuthoidea i squid) 36.36 0.62 Octopus dofleim (octopus) 72.73 4.63 Crangonidae (shrimp) 9.09 0.01 Pagurid (hermit crab) 9.09 0.01 Teleostei (unidentified fish) 4.5.45 0.33 Oncorhynchus sp. (salmon) 9.09 4.49 Gadidae (gadid fish) 9.09 0.49 Theragra chalcogramma (walleye pollock) 9.09 5.22 Atheresthes stomias (arrowtooth flounder) 63.64 67.21 Sebastes sp. (rockfish) 9.09 2.06 Pleuronectid (unknown flatfish) 18.18 0.86 Hippoglossoides elassodon ( flathead sole I 9.09 0.98 Fishery offal 9.09 12.27 Total prey weight (g) 24,017 Number of stomachs with food 11 Number of empty stomachs 2 Literature cited Bigelow, H. B., and W. C. Schroeder. 1948. Lancelets, cyclostomes and sharks. /;i Fishes of the western North Atlantic, part 1, p. 1-576. Sears Found. Mar Res.. Yale Univ., New Haven. CT. Bright, D. B. 1959. The occurrence and food of the sleeper shark, Somniosus pacificus, in a central Alaska bay. Copeia 1959 (11:76-77. Castro, Jose I. 1983. The sharks of North American waters. Texas A&M Univ Press, College Station. TX. 180 p. Crovetto, A., J. Lamilla, and G. Pequeno. 1992. Lissodelphis peronn. Lacepede 1804 (Delphinidae. Cetacea) within the stomach contents of a sleeping shark, Somniosus cf. pacificus. Bigelow and Schroeder 1944, in Chilean waters. Mar Mamm. Sci. 8(3):312-314. Gotshall, D. W., and T. Jow. 1965. Sleeper sharks, Somniosus pacificus. off Trinidad, California, with life history notes. Calif. Fish Game 51:294-298. NMFS (National Marine Fisheries Service). 1995. Status review of the United States Steller sea lion, Eumetopias juhatus, population. U.S. Dep. Commer, National Marine Mammal Laboratory, 7600 Sand Point Way NE. Seattle, WA 98115-0070, 45 p. Phillips, J. B. 1953. Sleeper shark, Somni(}sus pacificus. caught off Fort Bragg, California. Calif Fish Game 39( 1):147-149. Robinson. S. M. C, and E. B. Hartwick. 1983. Relationship between beak morphometries and live NOTE Yang and Page: Diet of Somniosus paaficus 409 wet weight of the giant Pacific octopus. Octopus dofleini martini (Wulker). Veliger 26(l):26-29. Sease, J. L., J. P.Lewis, D. C. McAllister, R. L. Merrick, and S. M. Mello. 1993. Aerial and ship-based surveys of Steller sea Hons, £;//?it'top!asj;/6a/us. in Southeast Alaska, the Gulf of Alaska, and Aleutian Islands during June and July 1992. U. S. Dep. Commer., NOAA Tech. Memo. NMFS-AFSC-17, .57 p. Tanaka, S., K. Yano, and T. Ichihara. 1982. Notes on a Pacific sleeper shark, Somniosus pad ftcus, from Suruga Bay. Japan. J. Fac. Mar Sci. Technol. Tokai Univ. 15:345-358. Wilimovsky, N. J. 1954. List of the fishes of Alaska. Stanford Ichthyol. Bull. 4(5):279-294. 410 Fishery Bulletin 97(2), 1999 Superintendent of Documents Publications Order Form *5178 I I I lliS, please send me the following publications: Subscriptions to Fishery Bulletin for $35.00 per year ($43.75 foreign) The total cost of my order is $ Prices include regular domestic postage and handling and are subject to change. (Company or Personal Name) (Please type or print) (Additional address/attention line! (Street address) (City, State, ZIP Code) (Daytime phone including area code) (Purchase Order No.) Charge your order. ITS EASY! VISA Please Choose Method of Payment: I I Check Payable to the Superintendent of Documents I I GPO Deposit Account [ I I VISA or MasterCard Account n 1 (Credit card expiration date) (Authorizing Signature) Mail To: Superintendent of Documents P.O. Box 371954, Pittsburgh, PA 15250-7954 To fax your orders (202) 512-2250 Thank you for your order! 3907 7 U.S. Department of Commerce Seattle, Washington Volume 97 Number 3 July 1999 Fishery Bulletin icud The National Marine Fisheries Service iNMFS) does not approve, recommend, or endorse any proprietary product or proprietary material mentioned in this publication No reference shall be made to NMFS, or to this pubhcation furnished by NMFS, in any advertising or sales promotion which would indicate or imply that NMFS approves, recommends, or endorses any proprietary product or proprietary material mentioned herein. oi which has as its purpose an intent to cause directly or indirectly the adver- tised product to be used or purchased because of this NMFS publication, Contents Articles 411-420 Austin, Herbert M., Daniel Scoles, and Allison J. Abell Morphometric separation of annual cohorts within mid-Atlantic bluefish, Pomatomus saltatrlx, using discriminant function analysis 421 -433 Bertignac, Michel, John Hampton, and Atilio L. Coan Jr. Estimates of exploitation rates for north Pacific albacore, Thunnus alalunga, from tagging data 434-448 Broadhurst, Matt K., Roger B. Larsen, Steven J. Kennelly, and Paul E. McShane Use and success of composite square-mesh codends in reducing bycatch and in improving size-selectivity of prawns in Gulf St. Vincent, South Australia 449-458 DeMartini, Edward E., and Boulderson B. Lau Morphometric criteria for estimating sexual maturity in two snappers, Etelis carbunculus and Pristipomoldes sleboldii 459-471 Franks, James S., James R. Warren, and Michael V. Buchanan Age and growth of cobia, Rachycentron canadum, from the northeastern Gulf of fviexico 472-481 Friedland, Kevin D., Jean-Denis Dutil, and Teresa Sadusky Growth patterns in postsmolts and the nature of the marine juvenile nursery for Atlantic salmon, Salmo salar 482-494 Hannah, Robert W. A new method for indexing spawning stock and recruitment in ocean shrimp, Pandalus jordani, and preliminary evidence for a stock-recruitment relationship 495-507 Labropoulou, Mary, Athanasios Machias, and Nikolaos Tsimenides Habitat selection and diet of juvenile red porgy, Pagrus, pagrus, (Linnaeus, 1758) 508-525 Lindeman, Kenyon C, and David B. Snyder Nearshore hardbottom fishes of southeast Florida and effects of habitat burial caused by dredging 526-541 Nates, Sergio P., and Darryl L. Felder Growth and maturation of the ghost shrimp Lepidophthalmus sinuensis Lemaitre and Rodngues, 1991 (Crustacea, Decapoda, Callianassidae), a burrowing pest in penaeid shrimp culture ponds 542-554 Perkins, Peter C, and Elizabeth F. Edwards Capture rate as a function of school size in pantropical spotted dolphins, Stenella attenuata, in the eastern tropical Pacific Ocean Fishery Bulletin 97(3), 1999 555-569 Rilling, Gene C, and Edward D. Houde Regional and temporal variability in growth and mortality of bay anchovy, Anchoa mitchilli, larvae in Chesapeake Bay 570-580 Roa, Ruben, Billy Ernst, and Fabian Tapia Estimation of size at sexual maturity: an evaluation of analytical and resampling procedures 581-590 Rooker, Jay R., Scott A. Holt, G. Joan Holt, and Lea A. Fuiman Spatial and temporal variability in growth, mortality, and recruitment potential of postsettlement red drum, Sclaenops ocellatus, in a subtropical estuary 591-603 Sigler, Michael F. Estimation of sableflsh, Anoplopoma fimbria, abundance off Alaska with an age-structured population model 604-616 Sladek Nowlis, Joshua, and Galium M. Roberts Fisheries benefits and optimal design of marine reserves 617-625 Somerton, David A., and Robert S. Otto Net efficiency of a survey trawl for snow crab, Chionoecetes opilio, and Tanner crab, C bairdi 626-635 Szedlmayer, Stephen T., and Joseph Conti Nursery habitats, growth rates, and seasonality of age-0 red snapper, Lutjanus campechanus. In the northeast Gulf of Mexico 636-645 Thomas, Ross, and Natalie A. Moltschaniwskyj Ontogenetic changes in size and shape of statoliths: implications for age and growth of the short-lived tropical squid Sepioteutliis lessoniana (Cephalopoda: Loliginidae) 646-660 Valle, Charles F., John W. O'Brien, and Kris B. Wiese Differential hatitat use by California halibut, Paralichthys californicus, barred sand bass, Paralabrax nebulifer, and other juvenile fishes in Alamitos Bay, California 661 -679 Van Eeckhaute, Lutgarde A. M., Stratis Gavaris, and Edward A. Trippel Movements of haddock, Melanogrammus aeglefinus, on eastern Georges Bank determined from a population model incorporating temporal and spatial detail 680-689 Woodbury, David Reduction of growth in otoliths of widow and yellowtall rockfish (Sebastes entomelas and 5. flavidus) during the 1983 El Nino 690-701 Xiao, Yongshun General age- and time-dependent growth models for animals 702-712 Ye, Yimin, and Hussain M. A. Mohammed An analysis of variation in catchability of green tiger prawn, Penaeus semisulcatus. In waters off Kuwait Notes 713-716 Griffith, Jason N., Andrew P. Hendry, and Thomas P. Quinn Straying of adult sockeye salmon, Oncorhynchus nerl400 mm) bluefish did not classify fish by probable spawning cohort. It did, how- ever separate the yearling fish (200- 400 mm) by year class rather than geo- graphic or seasonal spawning. Older fish (>400 mm) showed less separation because multiple (2-10) year classes were present. The DNA studies have revealed genetic homogeneity among these fish. This finding suggests that the morphological characteristics are phenotypically plastic and are influ- enced each year by the physical envi- ronment during spawning and the early juvenile stages. Morphometric separation of annual cohorts within mid-Atlantic bluefish, Pomatomus saltatrix, using discriminant function analysis* Herbert M. Austin Daniel Scoles Allison J. Abell School ol Marine Science Virginia Institute of Marine Science College of William and Mary Gloucester Point, Virginia 23062 Email address (for H M Austin) Haustin a'vims edu Manuscript accepted 14 August 1998. Fish. Bull. 97:411-420 (1999). The question of bluefish (Pomatomus saltatrix) stock composition along the east coast of the United States has been of considerable interest to fisheries scientists for 30 years and has recently (since 1986) become the focus of discussion by management agencies, primarily the Atlantic States Marine Fisheries Commis- sion (ASMFC) and Mid-Atlantic Fisheries Management Council (MAFMC) (Anonymous, 1989). The problems of managing fisheries when the stock composition and boundaries are unclear have ham- pered effective management of weakfish, Cynoscion regalis. sum- mer flounder, Paralichthys den- tatus, surf clam, Spisula solidis- sima. and striped bass, Morone saxatilis by the ASMFC or MAFMC (or both) because current manage- ment practices manage by unit (ge- netic) stock. Interstate manage- ment is further hampered when a unit stock exhibits differential spa- tial reproduction and migration pat- terns (e.g. weakfish, Scoles, 1990). The stock structure of the blue- fish (Pomatomus saltatrix) in the western mid-Atlantic region is not well understood. Lund (1961) re- corded meristic counts of gill rak- ers along the first bralnchial arch of young fish and suggested that six separate stocks of bluefish occur along the western north Atlantic coast of the United States. He later suggested the occurrence of several races in this region on the basis of results of two mark-recapture stud- ies (Beaumariage and Wittich, 1966; Lund and Maltezos, 1970). These results were not supportive of Lund's original six stock concept because they suggested that there might be a "Florida" or "South At- lantic Bight" and a "northern" or "mid-Atlantic Bight" stock. Subse- quent analyses of temporal and spa- tial distributions of bluefish in the ichthyoplankton, and spawning times, suggested two north Atlan- tic stocks (Norcross et al, 1974; Kendall and Walford, 1979), in ad- dition to a south Atlantic or Florida Stock (Collins and Stender, 1987). Chiarella and Conover ( 1990), on the other hand, using back calcula- tions of scales, demonstrated that most spawning in the north mid- Atlantic Bight (MAB ) occurs dur- ing mid-summer (July) and is com- posed of spring-spawned fish, both of which suggest a single stock. They also found from back calcula- tions that most yearling fish (260 mm) collected in waters along Long ' Contribution 2143 from the Virginia Insti- tute of Marine Science, College of William and Mary, Gloucester Point, VA. 412 Fishery Bulletin 97(3), 1999 Island during 1986-87 were spring-spawned. Later, McBride and Conover (1991), looking at young-of- the-year bluefish in the New York Bight during the summers of 1987 and 1988, found two discrete size groups by late summer (150 mm, 75 mm). Their otolith analyses confirmed that the fish represented both spring- and summer-spawned cohorts. More recent larval studies (Smith etal., 1994; Hare and Cowen, 1993) suggest an alternate hypothesis, that of a continuous wave of spawning by a single stock from off Hatteras in April-May to off Cape Cod or Block Island in June-July with two survival events, one in spring and one later in the summer as a result of oceanographic conditions. These survival events may have led to the previous hypothesis of two distinct spawning events. At any rate, the ge- netic analyses of Graves et al. (1993), using mtDNA, have shown that progeny from both the spring and summer spawning were of the same stock, and that mid-Atlantic Bight bluefish compose a single genetic stock. The various hypotheses have been revisited by Juanes et al. ( 1996) in a review of global bluefish early life history. A characterization of the seasonal movement and spawning of what was then (1977) considered the north Atlantic stocks was summarized by Wilk (1977). Before the advent of routine genetic testing, Wilk (1977) conducted a morphometric analysis of yearling fish from the North Carolina sounds and Middle Atlantic Bight to test the hypothesis that two stocks of bluefish occurred in the mid-Atlantic Bight. The preliminary data, results, and manuscript were lost in the 1984 NMFS/NOAA Sandy Hook Marine Laboratory fire. Wilk did find morphometric differ- ences that were statistically significant (Wilk'). If these two geographically separate spawnings are by the same stock, as demonstrated by Graves et al. ( 1993 ), but exhibit morphological differences as sug- gested by Wilk, perhaps due to environmental phe- notypic plasticity, then a potentially valuable tool for management exists, particularly if growth, and re- cruitment or harvest pressure (or both) are differ- ent. With this possibility in mind, we conducted a morphometric analysis of bluefish collected in the mid-Atlantic Bight of the U.S. east coast. April 1990 from several locations between eastern Long Island, New York, and Beaufort, North Caro- lina. Samples were pooled on the basis of geography and date-year of collection (Tables 1 and 2 ). Most were collected from pound and gill nets, but several small fish were taken by 10.8-m (30-ft) otter trawl. The majority of fish over 600 mm TL were collected by hook and line tournament fishermen. Twenty two morphometric measurements were recorded from the left side of the fish with a meter stick or dial calipers to the nearest millimeter. Names of morphometric variables and abbreviations are provided in Table 3. Scales were removed from un- der the pectoral fin of each fish, mounted on acetate sheets, and ages were determined with a microfilm reader according to the techniques of Hill and Loesch.- Samples were classified by using stepwise linear discriminant function analysis (DFA) (Fisher, 1936) with SPSS software program (Norusis, 1985). An excellent introduction to the statistics of discrimi- nant analysis is presented by Klecka ( 1989) in which all assumptions and shortcomings of the methods are discussed. Allometric growth can cause bias, and al- though it is recognized that it is impossible to re- move all allometric bias, Riest ( 1985), in a review of transformation methods, has offered Thorpe's ( 1975) as among the best in this situation. Schaefer (1990) and Scoles ( 1990) also used this technique and found it satisfactory for removing size effects during mor- phometric analyses of tuna and weakfish. Consequently, all measurements were transformed following equations taken from Thorpe ( 1975) where Y, = 10' y;=iog,„i^-6(iog„,z, -iog,„f) where F = the adjusted variable of the jth specimen; Y = the variable to be transformed of the (th specimen; b = the allometric coefficient; X^ - a standard measure of size of the ;th specimen for which fork length was used; _ and X= the grand mean of standard lengths. Methods A total of 1386 bluefish, ranging in size from 93 to 888 mm TL, were collected from April 1987 through 1 Wilk, S. 1989. Sandy Hook Marine Laboratory, National Ma- rine Fisheries Service, NOAA, Sandy Hook, NJ. Personal commun. A third equation, combining the first two provides Logi Y./Y, = 6Logi X./X. ^ Hill,B.,and J. Loesch. 1989. Striped bass research in Virginia: characterization of Virginia commercial fisheries. Annual Re- port 88-89, 22 p. Virginia Institute of Marine Science, P.O. 1346, Gloucester Point. VA 23062. Austin et al.: Morphometnc separation of annual cohorts within mid-Atlantic Pomatomus saltatnx 413 Table 1 Bluefish collection data: date, location, number of fish collected, and gear used in collection. L.I.= Long Island; L.I.S. = Long | Island Sound. Date Location n Gear 10 Mar 1987 Hatteras, NC 8 Gill net 30 Apr 1987 Aberdeen Creek, VA 10 Gill net 30 Apr 1987 Mobjack Bay, VA 21 Gill net 17 May 1987 Chesapeake Bay, VA 32 Hook and line 14 Jul 1987 Hatteras, NC 16 Seine 8 Jul 1987 Mobjack, VA 12 Gill net 2 Oct 1987 New Jersey 11 Trawl 6 Nov 1987 Chesapeake Bay, V 44 Hook and line 12 Apr 1988 Norfolk, VA 20 Gill Net 19 Apr 1988 Rappahannock R. 19 Pound net 1 May 1988 York River, VA 10 Pound net 12 May 1988 York River. VA 1 Pound net 16 Jun 1988 Ches. Bay, VA 58 Hook and line 18 Jul 1988 Pt. Lookout, L.L 50 Gill net 19 Jul 1988 Peconic Bay. L.I. 24 Gill net 19 Jul 1988 Montauk, L.I. 71 Hook and line 2.5 Jul 1988 York River, VA 26 Pound net 28 Jul 1988 Hatteras, NC 85 Pound net 3 Aug 1988 York River, VA 8 Pound net .5 Aug 1988 York River, VA 76 Pound net 5 Aug 1988 L.LS., CT 15 Trawl 9 Sep 1988 L.I.S.,CT 171 Trawl 9 Sep. 1988 Potomac R. 69 Hook and line 4 Apr 1989 Pamlico Sound 81 Various 4 Apr 1989 Oregon Inlet. NC 169 Various \5 June 1989 Reedville, VA 86 Hook and line 25 July 1989 New York. NY 51 Various 15 Aug 1989 York River. VA 103 Pound net 9 Sep 1989 Hatteras. NC 42 Pound net which more clearly shows Y^ is an estimate of the average Y^ for an individual of fork length X^. Following transformation, each variable was re- gressed against fork length (FL). The slope of each transformed variable on FL was zero or insignificant in all cases; therefore effects of allometry were dis- regarded. The results were plotted for visual inspec- tion of outliers which were removed before subse- quent analyses if they were outside the range of bio- logical possibility, and thus suggested measurement error. Consequently, two subgroups of data were de- veloped. The first included bluefish between 200 and 400 mm fork length (yearlings), to remove young-of- the-year from the data which are in the stage of growth most likely to show allometry and to dupli- cate the size range used by Wilk and Walford in 1964 (Wilk, 19771. The second group included bluefish greater than 400 mm fork length. The linear discriminant function used here is of the form D = BiXj + B.2X., + B3X3 + B„X„+C and is similar to a multiple linear regression where D = the discriminant function that char- acterizes each reference group; X 's - theindependent variables (individual measurements) selected at in a step- wise fashion; and 414 Fishery Bulletin 97(3), 1999 Table 2 Discriminant function analysis sample groupings by location, year, and size category. Ches Bay = Chesapeake Bay. ID Sample size and len gth range (mm, in parentheses) Group Year number 200-400 mm >400 mm Summer Hatteras 1987 1 14 (303- -382) 2(417- -441) Summer Ches Bay 1987 2 10(258- -316) 37(448- -884) Spring Hatteras 1987 3 — 8 (684-765) Spring Ches Bay 1987 4 15 (325- -385) 30 (801- -852) Summer Ches Bay 1988 5 11 (225- -382) 106(570- -830) Summer Long Isl. 1988 6 7(219- -395) 174 (555- -811) Spring VA Coast 1988 7 — 13 (471- -731) Spring Ches Bay 1988 8 — 26 (471- -757) Summer Ches Bay 1988 9 104 (245- -2911 — Summer Hatteras 1988 10 85 (235- -321) — Spring Hatteras 1989 11 42(357- -398) 101 (477- -784) Summer Ches Bay 1989 12 128(215- -352) 51 (405- -525) Summer Long Isl. 1989 13 — 44 (418- -734) Summer Hatteras 1989 14 — 29(412- -516) Spring PamHco 1989 15 71(357- -398) 7(402- -715) Spring Ches Bay 1990 16 524(215- 26' -398) 658 (402- 30' -884) ' Length data lost. Table 3 List of morphometric measurements on bluefish. Abbre- viations and name of ifariable are those used in text. Abbreviation Description PMX Premaxilla MAX Maxilla lOB Interorbital POB Postorbital POP Preoperculum ("cheek") OPC Operculum HDP Head depth PCO Pectoral fin origin PCI Pectoral fin insertion PLO Pelvic fin origin PL! Pelvic fin insertion VNT Vent AFO Anal fin origin API Anal fin insertion DIO First dorsal fin origin Dll First dorsal fin insertion D20 Second dorsal fin origin D2I Second dorsal fin insertion GTH Girth TOL Total length FKL Fork length STL Standard length DSP First and second dorsal space B^^'s= the coefficients or "unstandardized fianction coefficients"; and C = a constant. Initially we intended to follow the 1960s methods of Wilk (1977) by assigning a spring-southern and summer-northern a priori reference designation. We were unable however, to collect a reference group of spring- and summer-spawned yearling bluefish. From the length frequencies of the yearling fish (211- 382 mm) that we collected, it appeared that all were "spring-spawned" (Chiarella and Conover, 1990). As a result we made no effort a priori to separate "spring- spawned" from "summer-spawned" yearling fish on the basis of scale back calculations as Chiarella and Conover { 1990) had done but assumed all were spring-spawned. A priori assignment of ref- erence or learning groups is often used to determine the discriminant function which is then used to clas- sify the individuals of known origin to one or the other reference groups. If, however, more than two stocks are present, the individuals of the third stock will be "force fitted" into one of the reference groups. Rather than forcing a pnoW assignments of two groups, and to identify possible additional stocks (from the south IMcBride et al. 1993 1) or morphometrically distinct Austin et aL: Morphometric separation of annual cohorts within mid-Atlantic Pomatomus saltatnx 415 ra o 'c o c 400 mm; however, the eigenvalues (Tables 5 and 6) show that only the first two functions are important in each case. The discriminant scores and centroids from functions 1 and 2 were plotted against each other to develop a graphic representation of the relationship among gi-oups (Figs. 1 and 2). Wilk's 1960's ( 1977) analysis found that the inter- dorsal space was a discriminating character. To pro- vide a comparison, we computed an "inter-dorsal Table 4 Summary of stepwise discriminant function analysis for 16 groups and 16 morphometric characters. Fish were 200- 400 mm FL. See Table 3 for definitions of variables. Minimum Step no. Variable entered Wilk's lambda P D P 1 PLI 0.44335 <0.0001 0.00764 0.8081 2 HDP 0.34477 <0.0001 0.23817 0.2317 3 MAX 0.25160 <0.0001 0.65660 <0.0001 4 DSP 0.16585 <0.0001 0.87455 0.0001 5 lOB 0.13113 <0.0001 1.20062 <0.0001 6 OPC 0.12388 <0.0001 1.29810 <0.0001 7 PCO 0.10184 <0.0001 1.44335 <0.0001 8 AFO 0.09450 <0.0001 1.47319 <0.0001 9 D2I 0.08292 <0.0001 1.55493 <0.0001 10 DIO 0.07776 <0.0001 1.62040 <0.0001 11 AFI 0.07325 <0.0001 1.67691 <0.0001 12 POB 0.06806 <0.0001 1.69120 <0.0001 13 POP 0.06544 <0.0001 1.70979 <0.0001 14 PMX 0.05928 <0.0001 1.72065 <0.0001 1.5 PLO 0.05329 <0.0001 1.72103 0.0001 16 Dll 0.04386 <0.0001 1.72130 0.0001 416 Fishery Bulletin 97(3), 1999 Table 5 Summary of first five canon cal discriminant functions . Fish were 200- -400 mm FL. Cumulative Canonical Wilk-s Chi Function Eigenvalue percent correlation lambda squared df P 1 2.29051 48.91 0.834324 0.15912 935.6 150 <0.0001 2 1.16710 73.84 0.733863 0.34483 541.6 126 <0.0001 3 0.54375 85.45 0.593486 0.53234 320.9 104 <0.0001 4 0.28451 91. .53 0.470633 0.68379 193.5 84 <0.0001 5 0.14586 94.64 0.356785 0.78354 124.2 66 <0.0001 Table 6 Summary of the first seven canonical discriminant functions. Fish were >400 mm FL. Cumulative Canonical Wilks Chi Function Eigenvalue percent correlation lambda squared df P 1 1.4858 42.4 0.77312 0.1880 1072.9 180 <0.0001 2 0.8153 65.6 0.67016 0.3413 690.1 154 <0.0001 3 0.4462 78.4 0.55546 0.4936 453.2 130 <0.0001 4 0.2595 85.8 0.45387 0.6217 305.1 108 <0.0001 5 0.1682 90.6 0.37945 0.7263 205.3 88 <0.0001 6 0.0900 93.1 0.27740 0.7917 149.9 70 <0.0001 7 0.0769 95.4 0.26722 0.8526 102.4 54 <0.0001 Table 7 Summary of stepwise discriminant function analysis for | 16 groups and 16 morphometric characters. Fish were >400 mm FL. See Table 3 for definitions of variables. Minimum Step Variable Wilk's no. entered lambda P D P 1 Dll 0.7177 <0.0001 0.0007 0.9364 2 POP 0.6367 <0.0001 0.0288 0.8088 3 PLO 0.4628 <0.0001 0.2543 0.0520 4 DSP 0.4326 <0.0001 0.4455 0.1638 5 HDP 0.2946 <0.0001 0.6365 0.0008 6 AFO 0.2645 <0.0001 0.7851 0.0237 7 PCO 0.2483 <0.0001 0.8786 0.0225 8 PLI 0.1813 <0.0001 1.0243 <0.0001 9 DIO 0.16.53 <0.0001 1.2288 <0.0001 10 POB 0.1323 <0.0001 1.2774 <0.0001 11 JOB 0.1090 <0.0001 1.3481 <0.0001 12 D2I 0.1025 <0.0001 1.4211 <0.0001 13 OPC 0.0994 <0.0001 1.4769 <0.0001 14 MAX 0.0996 <0.0001 1.5093 <0.0001 15 PMX 0.0776 <0.0001 1.5274 <0.0001 16 AFI 0.0756 <0.0001 1.5402 <0.0001 space" term ( DSP), as the difference between the first dorsal insertion (DID less the second dorsal origin (D20). When the analysis was run with this new variable (DSP), it was selected as an important term in the functions (Tables 4 and 9). The centroids for yearling fish do not fall into two classes, which would have suggested either a geo- graphic north-south or temporal spring-summer spawned classification (Fig. 1); rather they fall into three clusters of cohorts by year class, regardless of geography of collection site (Table 10). For example, collections during 1987 at Hatteras, North Carolina, and Chesapeake Bay, Virginia (the 1986 year class), are separate and distinct from both the 1988 and 1989 collections (1987 and 1988 year classes) and also from the Hatteras and Chesapeake Bay collections. Further, the 1989 collections ( 1988 year class ) were clas- sified into two groups, one of which was the same as the 1988 collection. The 1990 col lection ( 1989 year class ) fell in between 1988 and 1989, overlapping both. The centroids of the large fish (>400 mm), a mix- ture of up to 10 year classes in a sample, showed little geographic, temporal, or year-class classifica- tion (Fig. 2). Austin et al : Morphometnc separation of annual cohorts within mid-Atlantic Pomatomus saltalnx 417 R n Bluefisti >400mm /i\ 1987 Centroids S 1988-89 Centroids * 1990 Centroid cv 4.0- c q 1 2.0- c c I .0- S -2.0 - c " -4.0- -6.0- Spring/Summer Hatteras, Chesapeake Bay.Long Island Summer 1988-1989 -• ■ .^ Chesapeake Bay 1990 / a/ \ ^, Spring/Summer Chesapeake Bay 1987 -6,0 -4.0 -2.0 .0 2.0 4.0 6.0 Canonical discriminant function 1 Figure 2 Scatterplot of canonical discriminant functions 1 and 2 for bluefish >400mm showing locations of group centroids for 1987-90 by geographic area of collection. Discussion Several authors (Nyman and Conover, 1988; McBride, 1989; Simpson et al.. 1990) working in the Long Is- land region and analyzing length-frequency data, reported exclusively spring-spawned young-of-the- year during 1985-86. Both spring-spawned and some late arriving summer-spawned young-of-the-year were collected there in 1987, and then a shift occurred to predominantly summer-spawned fish during the summer-fall of 1988. Because ratios continued to be similar for the year- ling fish in our collections and because length fre- quencies were also similar, our 1987 samples from Hatteras and the Chesapeake Bay were apparently from the predominantly spring-spawned 1986 year class. Our 1988 samples, the 1987 year class, were also predominantly spring spawned but were classi- fied morphologically separate from the 1986 year class. Our 1989 samples (the 1988 year class) from Hatteras, Chesapeake Bay, and Long Island showed a split classification, some the same as our 1988 samples (the 1987 year class) and the rest, separate from both 1987 and 1988 year classes. The 1990 samples, from a dominant 1989 year class, collected in the Chesapeake Bay, overlapped slightly between the 1987 and 1988 year classes (Fig. 1). If the shift from spring-spawned to summer- spawned young-of-the-year noted by the above au- thors for Long Island holds for Chesapeake-Hatteras yearling fish, then the morphologically distinct 1986 year class from our 1987 collections was composed of spring-spawned fish. The 1988-90 collections of the 1987-89 year classes, may be a mix of spring- and summer-spawned fish but show no separation. As stated earlier, however, our length frequencies of yearling fish in all years ( 1987-90) suggest that our samples were all spring-spawned fish (Table 2). An alternate explanation is that the morphologically distinct yearling fish, representing separate year classes or annual cohorts, and spawned in differing environments each year are demonstrating environ- mental or phenotypic plasticity. That is, that their morphological characters are environmentally deter- mined and are different each year. 418 Fishery Bulletin 97(3), 1999 Table 8 Canonical discriminant function coefficients linear discrim- inant equation (Norusis. 1985): D = B„ + S,Ar, + B^., + S3X3 BJC^. See Table 3 for definitions of variables. Variable Unstandardized function coefficients 200-400 mm FL function >400 mm FL function 1 2 1 2 PMX -11.3805 -7.5218 -5.8925 -27.9831 MAX 1.2972 4.1059 8.0311 30.1133 lOB 11.3277 -0.4940 2.0227 24.8566 POB -6.1778 -18.1707 -5.7929 -45.9063 POP 6.1848 4.7921 -5.1122 25.7729 OPC -26.3362 6.4774 -2.5250 -29.3489 HDP -0.6669 29.7938 26.4669 17.4505 PCO 36.7537 -2.9036 1.8195 -20.8899 PLO 10.1294 -13.7243 20.1050 -4.4471 PLI 26.8615 31.4590 -57.1219 26.6134 .\F0 -31.2089 8.5720 42.9709 -2.0460 .\FI 31.6723 35.3519 -24.5871 12.8353 DIO 4.0019 -3.4074 -35.3567 -8.2396 Dll 12.8394 -22.4628 48.7193 -14.1872 DSP 0.2674 3.7165 1.3185 1.3386 D2I 32.9298 -6.5686 3.9964 39.1365 CONSTANT -212.7263 -118.6822 -45.9465 -78.7146 The larger fish in this study are a composite of at least ten year classes (2-11 years); therefore separa- tion or classification would be expected to be less precise. Indeed there was considerable overlap in classification of the larger fish (Table 10). From results of the DFA, and in light of the back calculations of Chiarella and Conover ( 1990) and the genetic data of Graves et al. ( 1993 ), it would appear that there is only one stock of bluefish in the mid- Atlantic Bight and that the morphometric differences among bluefish cohorts are a result of phenotypic plasticity. Because the environment at the time of spawning and juvenile development varies geo- graphically and interannually, so too will morpho- metric features. As the fish grow, their plastic mor- phological characters, expressed as an index of char- acter length versus fish length, become less and less reliable: when year classes mix, as in our >400 mm sample, they provide no discriminant characteriza- tion unless separated by individual year class. Although a single genetic stock, the two MAB co- horts (spring- and summer-spawned), may exhibit interannual differential recruitment success and sur- vival to yearlings. In addition! the reported differ- Table 9 Canonical discriminant function group centroid means. * = no collections from this size group Group 200-400 mm >400 mm Function 1 Function 2 Function 1 Function 2 1 -1.1747 -2.4628 2.2954 -4.1523 2 -1.8353 -4.8299 -0.7644 -2.2536 3 * -0.9446 0.5734 4 -1.4581 -2.4707 -0.4563 -2.1051 5 0.2588 1.6696 0.9954 0.6740 6 -0.4975 0.3665 1.3013 -0.0488 7 * 0.8351 -0.6780 8 * 1.4361 -0.1726 9 -1.4280 -0.0648 * 10 -1.0596 0.8962 * 11 -0.7187 1.0913 -0.0650 0.3479 12 2.2186 -0.3778 -1.1776 0.1391 13 * -1.5116 0.4816 14 2.7864 1.0578 -1.3068 1.6483 15 -0.2675 0.3093 -0.7993 0.5799 16 1.1706 -0.0615 -1.9122 0.4894 ences in growth, seasons and location of spawning, migration routes, and variations in fishing mortal- ity along the MAB can complicate management ef- forts if the two cohorts are managed as a single stock. Even for a single mid-Atlantic genetic stock, recruit- ment variations between spring and summer cohorts may be significant, year to year, and consideration should be given to monitoring the annual contribu- tion of the spring-spawned and summer-spawned cohorts as the stock may be healthier when the spring-spawned predominate for several years run- ning (Chiarella and Conover, 1990). The morphomet- ric separation by year class provides evidence that the environment likely affects bluefish size and shape; this may prove to be a useful tool in separat- ing yearling bluefish by the geographic area in which they spent their first year of development. Conclusions Previous larval and tagging studies neither support nor refute the existence of one or more MAB blue- fish stocks, one that spawns just south of Hatteras, North Carolina, in the spring and the other off New England in the summer. The mtDNA analysis by Graves et al. ( 1993) suggests that there is one mid- Atlantic stock which from our results produces sev- eral environmentally induced morphotypes. Austin et al : Morphometric separation of annual cohorts within mid-Atlantic Pomatomus saltatnx 419 Table 10 Classification resu HA = Hatteras; CE Its by geographic area = Chesapeake Bay; h and year for fish 200-400 mm FL and fish >400 mm = Long Island; PS = Pamlico Sound, NC. FL. SU = Summer; SP = Spring; Group Group number Predicted group membership (%) n 1 2 3 4 5 6 7 8 9 10 11 12 13 14 15 16 200-400 mm FL SU/HA/87 1 14 64 7 7 7 14 SU/CB/87 2 10 100 SP/CB/87 4 15 7 47 27 13 7 SU/CB/88 5 11 46 27 9 18 SU/LI/88 6 7 29 14 29 14 14 SU/CB/88 9 104 1 1 76 14 3 3 2 SU/HA/88 10 85 1 19 61 7 1 11 SP/HA/89 11 42 2 36 55 2 5 SU/CB/89 12 128 1 5 2 2 79 2 3 6 SU/HA/89 14 11 64 27 9 SP/PS/89 15 71 13 16 1 1 69 SU/CB/90 16 26 4 4 4 15 4 69 >400 mm FL SU/HA/87 1 2 100 SU/CB/87 2 37 73 14 8 3 3 SP/HA/87 3 8 88 12 SP/CB/87 4 30 13 70 7 7 3 SU/CB/88 5 106 1 53 26 2 9 4 1 1 3 SU/LI/88 6 174 1 8 76 1 1 12 1 1 1 SU/VC/88 7 13 54 39 7 SP/CB/88 8 26 4 4 8 27 8 35 15 SP/HA/89 11 101 2 2 5 1 3 66 7 6 5 4 SP/CB/89 12 51 2 4 4 6 2 33 35 6 6 2 SU/LI/89 13 44 2 2 2 41 5 41 5 2 SU/HAy89 14 29 10 10 4 69 7 SU/PS/89 15 7 14 86 SU/CB/90 16 30 7 3 11 12 7 7 47 Acknowledgments Literature cited We wish to acknowledge the field assistance of Ana Beardsley and Jan McDowell, as well as the officials of the various Virginia bluefish tournaments that made samples available to us. Critical review was provided by John Graves and Jack Musick of VIMS. The most helpful comments of an anonymous re- viewer strengthened the manuscript and we are most grateful to this anonymous colleague. The work was supported by a grant to the Virginia Institute of Marine Science from the U. S. Fish and Wildlife Ser- vice, through the Virginia Marine Resources Com- mission, grant n number F-60-R. Anonymous. 1 989. Bluefish Fisheries Management Plan. Atlantic States Marine Fisheries Commission, Wash., D.C.. 54 p. Beaumariage, D., and A. Wittich. 1966. Returns from the 1964 Schlitz tagging program, Florida. Board Cons. Tech. Ser. 47, 50 p. Florida Dep. Natural Resources, St. Petersburg, FL. Chiarella, L. A., and D. O. Conover. 1990. Spawning season and first-year growth of adult blue- fish from the New York Bight. Trans. Am. Fish. Soc. 119:45.5-462, Collins, M. R. and B. W. Stender. 1988. Larval king mackerel iScomberomorus cavalla), Spanish mackerel 'S. maculatus), and bluefish iPomatomus 420 Fishen/ Bulletin 97(3), 1999 saltatrix) off the southeast coast of the United States, 1973- 1980. Bull. Mar. Sci.. 41:822-834. Fisher, R. 1936. The use of multiple measurements in taxonomic problems. Ann. -Eugenics 7:179-188. Graves, J. E., J. R. McDowell, A. M. Beardsley, and D. R. Scoles. 1993. Population genetic structure of the bluefish, Pomatomus saltatrix. in Atlantic coastal waters. Fish. Bull. 90:469-475. Hare, J. A., and R. K. Cowen. 1993. Ecological and evolutionary implications of the lar- val transport and reproductive strategy of bluefish, Pomatomus saltatrix. Mar Ecol. Prog. Ser. 98:1-16. Juanes, F., J. A. Hare, and A. G. Miskiewicz. 1996. Comparing early life history strategies of Pomatomus saltatrix: a global approach. Mar. Freshwater Res. 47:365-379. Kendall, A., and L. Walford. 1979. Sources and distribution of bluefish, Pomatomus saltatrix, larvae and juveniles off the east coast of the United States. Fish. Bull. 77(11:213-227. Kleeka, W. R. 1989. Discriminant analysis. Series: Quantitative applica- tions in the social sciences. Sage Publ., Inc., Thousand Oaks, CA, 70 p. Lund, W. 1961. A racial investigation of bluefish, Pomatomus saltatrix of the Atlantic Coast of North America. Bol. del Inst'o. Oceanog. Venezuela 1(11:73-129. Lund, W, and G. Maltezos. 1970. Movements and migrations of the bluefish, Pomato- mus saltatrix tagged in waters of New York and southern New England. Trans. Am. Fish. Soc, 99(4):719-725. McBride, R. S. 1989. Comparative growth and abundance of spring- ver- sus summer- spawned young-of-the-year bluefish, Poma- tomus saltatrix. recruiting to the New York Bight. M.S. thesis submitted to the State Univ. of New York at Stony Brook, NY, 56 p. McBride, R. S., and D. O. Conover. 1990. Recruitment of young-of-the-year bluefish Pomato- mus saltatrix to the New York Bight: variation in abun- dance and growth of spring- and summer-spawned cohorts. Man Ecol. Prog. Sen 78:205-216. McBride, R. S., J. L. Ross, and D. O. Conover. 1993. Recruitment of bluefish Pomatomus saltatrix to es- tuaries of the U.S. South Atlantic Bight. Fish. Bull. 91:389-395. Norcross, J. J., S. L. Richardson, W. Massman, and E. Joseph. 1974. Development of young bluefish, Pomatomus saltatrix and distribution of eggs and young m Virginia waters. Trans. Am. Fish. Soc. 103(3):477-497. Norusis, M. 1985. SPSS advanced users guide. SPSS Inc., Chicago, IL, 474 p. Nyman, R. M., and D. O. Conover. 1988. The relation between spawning season and the re- cruitment of young-of-the-year bluefish. Pomatomus saltatrix. to New York. Fish. Bull. 86(2):237-250. Ricst, J. D. 1985. An empirical evaluation of several univariate methods that adjust for size variation in morphometric data. Can. J. Zool. 63:1429-1439. Schacfer, K. M. 1990. Geographic variation in morphometric characters and gill-raker counts of yellowfin tuna, Thunnus albacares, from the Pacific Ocean. Fish. Bull. 89:289-297. Scoles, D. R. 1990. Stock identification of weakfish. Cynoscion regalis. by discriminant function analysis of morphometric char- acteristics. M.S. thesis submitted to the College of Will- iam and Mary, School of Marine Science, Virginia Institute of Marine Science, Gloucester Point, VA 23062, 51 p. Simpson, D. G., P. T. Howell, and M. W. Johnson. 1990. Assessment of bluefish in Long Island Sound with reference to the coastwide stock. Section 2, job 6 m A study of marine recreational fisheries in Connecticut., p 83- 104. Connecticut Dep. Environ. Cons., Hartford. CT. Smith, W., P. Berrien, and T. Pottoff. 1994. Spawning patterns of bluefish, Pomatomus saltatrix, in the northeast continental shelf ecosystem. Bull. Mar. Sci. 54(1):8-16. Thorpe, R. S. 1975. Quantitative handling of characters useful in snake systematics with particular reference to intraspecific varia- tion in the ringed snake, Natrix natrix. Biol. J. Linn. Soc. 7:27-43. Wilk, S. 1977. Biological and fisheries data on hluedsh. Pomatomus saltatrix. U.S. Dep. Commer., NMFS/NOAA, Tech. Ser. Rep. 11, 55 p. 421 Abstract.— North Pacific albacore tag- ging data from a tag-release program conducted from 1971 to 1989 were ana- lyzed to obtain estimates of exploitation rates and related parameters. The ma- jor albacore fishing fleets in the North Pacific, the U.S. baitboat, Japan baitboat, troll and longline fleets were used in the analysis. Another category of fleet ("other") that combined remain- ing miscellaneous recapture sources was also used. Tag-attrition models in- corporating variable availability of tagged albacore to the various fleets, seasonal catchability, and multiyear effects on catchability were developed and applied. The incorporation of all three effects was found to improve model fit significantly. If exploitation of the tagged population is representa- tive of the North Pacific albacore popu- lation as a whole and if tag reporting rates were high, the results would sug- gest that the exploitation rate has been less than 10<7r per year since the early 1970s. However, a deficit of returns from the troll fleet in comparison with its catch suggested that the pattern of exploitation of the tagged population, by this fleet at least, was different from that for the untagged albacore popula- tion. After compensating for assumed depressed availability of tagged alba- core to the troll fleet, annual exploita- tion rates were estimated to have de- clined from a high of 40*?^ in the niid- 1970s to <10'7r since the early 1980s. Estimates of exploitation rates for north Pacific albacore, Thunnus alalunga, from tagging data Michel Bertignac John Hampton Secretariat of the Pacific Community B P D5 98848 Noumea Cedex, New Caledonia Email address (for M Bertignac)) mictielbeiaispc org nc Atilio L. Coan Jr. Southwest Fisheries Science Center PO Box 271 La Jolla, California 92038 Manuscript accepted 17 August 1998. Fish. Bufl. 97:421-433 ( 1999). Albacore tuna, Thunnus alalunga, occur in the tropical and subtropi- cal waters of the Pacific Ocean, where they comprise separate stocks north and south of the equa- tor (Murray, 1994). In the north Pacific, albacore spawn predomi- nantly in the spring and summer over a wide area bounded approxi- mately by 10°-30°N (Nishikawa et al., 1985). Juvenile albacore appear in the North Pacific Transition Zone (35°-45°N), where they are ex- ploited by the surface fishery, com- prising mainly Japanese and U.S. baitboat, troll, and until the early 1990s, Japanese and Taiwanese driftnet fleets (Fig. 1). Adult alba- core are found at lower latitudes, where they are caught by the sub- surface longline fishery, comprising Japanese, Korean, and Taiwanese fleets (Fig. 1). Longline catches have been fairly stable at 10,000-20,000 t annually. Total surface fishery catches (which have been domi- nated by Japanese baitboat catches ) have declined from a peak of around 100,000 t in the mid-1970s to 20,000-50,000 1 in recent years (Liu and Bartoo, 19951. Since 1971, 23,780 albacore have been tagged in the North Pacific in a program conducted jointly by the U.S. National Marine Fisheries Ser- vice (NMFSl and the American Fishermen's Research Foundation. The tag recoveries have been used in studies of albacore movements (Laurs and Lynn, 1977), growth (Laurs and Wetherall, 1981), and fishery interaction (Kleiber and Baker, 1987). However, to date no formal analysis of the data has been undertaken to estimate exploitation rates and related parameters. In this paper, we develop tag-attrition models (Kleiber et al., 1987), incor- porating different structural as- sumptions concerning availability and catchability of the tagged popu- lation. The models are fitted to the albacore tagging data to obtain es- timates of the instantaneous rate of natural mortality and fleet-specific catchability coefficients. These es- timates are used to construct time series of estimated exploitation rates. The relative merits of the alternative models and applicability of the esti- mates to the North Pacific albacore population in general are discussed. Materials and methods Tagging data Albacore tagging occurred from 1971 to 1996, mainly during April-Sep- 422 Fishery Bulletin 97(3), 1999 130E 150E 170E 170W 150W 130W 110W Longline ~^~~\ r Troll Longline U.S. baitboat It I L J I I I L 130E 150E 170E 170W 150W 130W now Figure 1 Schematic diagram of geographic distributions of major North Pacific albaeore fishing fleets. 3,000 -, 1970 1975 1980 1985 Year 1990 1995 Figure 2 North Pacific albaeore tag releases, by year. i tember, with 97% of the 23,780 releases occurring prior to 1987 (Fig. 2). Tagging was carried out aboard U.S. troll vessels and baitboats by NMFS technicians and commercial fishermen trained in tagging. Tag- ging methods are described by Laurs et al. (1976) and Laurs and Wetherall (1981). Only albaeore judged to be in very good condition were selected for tagging; therefore mortality due to tagging is believed to have been negligible. Tagged albaeore were re- leased over a wide area of the eastern and central North Pacific; the largest numbers of releases took place in coastal waters adjacent to North America at 30°-50°N (Fig. 3). Fewer albaeore were tagged west of 180°, which is the main operational area of the Japanese baitboat fleet. The size distribution of tagged albaeore is t3T3ical of surface fishery catches (Fig. 4). As of 14 August 1997, 1302 tagged albaeore (5.5% of the releases) had been recaptured and the tags returned to NMFS. Recaptures have been recorded from a wide range of national fleets and gear types, but the largest numbers of recaptures (68% of the total) were made by U.S. and Japanese baitboats and U.S. troll boats (Table 1). Recaptures occurred throughout the North Pacific, but most recaptures were concentrated in coastal waters adjacent to North America (mainly U.S. troll and baitboats) and west of 180° (Japanese baitboats) (Fig. 5). Bertignac et al.: Estimates of exploitation rates for Thunnus alalunga from tagging data 423 130E 150E 170E 170W 150W 130W now i \ • • • • • • 01 o Z o 130E 150E 170E 170W 150W 130W 110W Figure 3 Geographical distribution of North Pacific albacore releases, 1971-89 Population dynamics model for tagged albacore We used a tag-attrition model (e.g., Seber, 1973; Kleiber et al., 1987) to describe the dynamics of tagged albacore (hereafter referred to simply as "tags"). The model may be represented by ':,/ =«,(l-a)i37exp ijf Z,^,./+^+ w l-exp(-X,i^,/-^-V/)j. (1 where 'jf a - P = Y = the estimated number of tags from re- lease group ( recaptured by fleet f in time period J and returned with com- plete information (recapture time and fleet); the number of tags released in release group ;; the proportion of type- 1 tag losses; the proportion of recaptured tags that are returned; the proportion of returned tags having complete information; the instantaneous rate of fishing mor- tality on release group ; in time period j by fleet f; M J k = the instantaneous rate of natural mortality; = the instantaneous rate of type-2 tag loss; - an index for release group; = an index for time period; = an index for time periods prior to^; and = an index for fleet. Data stratification In order to allow seasonal variability in tag returns to be modeled, tag releases and returns were grouped by the quarter (Jan-Mar, Apr-Jun, Jul-Sep, and Oct-Dec) and year of release and recapture. This resulted in 48 release groups (Table 2). Small num- bers of releases in 1994 and 1995, from which no re- captures have so far been reported, were not included in the analysis. We defined five fleets: 1) U.S. baitboat; 2) Japanese baitboat; 3) U.S. and Canadian troll (re- ferred to as the troll fleet); 4) Japanese, Korean, and Taiwanese longline (referred to as the longline fleet); and 5) others (consisting of recapture sources not oth- erwise defined). This definition was designed to group fleets having similar characteristics (fishing methods, seasonal distributions of effort, and sizes of albacore caught). The observed tag returns, r^^ were thus grouped by tag release group ii), quarter and year of recapture (j), and recapture fleet if). Tag returns to the end of 1992 were considered in the analysis. 424 Fishery Bulletin 97(3), 1999 2,000 1,500 S 1,000 6 500 - [ n 1 1 I Pf 35 40 44 48 52 56 60 64 68 72 76 80 84 Fork length (cm) n 1 1 1 1 1 92 96 Figure 4 Size distribution of North Pacific albacore tag releases, 1971-89. Table 1 Number of albacore tag returns in the North Pacific 1971- 1992, by recapt jre fleet. Vessel or Number of gear type Nationality returns % Baitboat 715 56.3 U.S. 403 Japan 312 Troll 148 11.7 U.S. 146 Canada 2 Sport 137 10.8 U.S. 135 Japan 2 Longline 79 6.2 Japan 61 Korea 15 Taiwan 2 Unknown 1 Purse seine 39 3.1 U.S. 34 Japan 4 Mexico 1 Drift gill net 29 2.3 U.S. 1 Japan 24 Canada 1 Taiwan 3 Handline 5 0.4 U.S. 5 Unknown 117 9.2 U.S. 112 Japan 2 Unknown 3 Total 1269 100.0 Nuisance parameters We assumed that tag shedding was the only source of type- 1 and type-2 tag loss and therefore used val- ues for a and t// of 0.12 and 0.0245 per quarter, re- spectively, based on analysis of double-tagging data (Laurs et al., 1976). There are no independent esti- mates of nonreporting of North Pacific albacore tags, although Kleiber and Baker (1987) stated that "nonreporting losses are small for the major fisher- ies that recovered the tags"; they assumed p = 0.9. We also made this assumption but tested the sensi- tivity of our results to values of /3in the range 0.1 to 1.0. The value of /( 0.87) was estimated directly from the tag-return data. Reparameterization of Fishing mortality To create sufficient degrees of freedom to enable a statistical estimation of parameters, the number of fishing mortality parameters to be estimated, F„y needs to be reduced. If we assume for the moment that F,jf is independent of release group /, F if may be reparameterized as a function of fishing effort, Ejf, and catchability, g,/ : 'jf (2) We may then propose models that constrain q . in some sensible way. Let S^ denote a matrix that con- tains, for each f, a season for each time period in- dexed hy j. We tested two seasonal schemes: no sea- sonal variation for any fleet and seasonal variation (Ql=Jan-Mar, Q2=Apr^un, Q3=Jul-Sep, Q4=Oct- Dec) for all fleets. Bertignac et al : Estimates of exploitation rates for Thunnus alalunga from tagging data 425 130E 150E 170E 170W 150W 130W 110W T o o CO o n yjy • • • • • • • # • • ••••• Albacore tag recoveries • 250 • 125 • 25 J L J L tn o Z o o 2 130E 150E 170E 170W 150W 130W now Figure 5 Geographical distribution of North Pacific albacore recoveries, 1971—92 Additionally, we tested constraints on Qji that speci- fied how catchability might vary over longer (multiyear) time periods. Let T,/ be a matrix that, for each f, maps time period / into a series of multiyear periods within which the seasonal pattern of catchability for that fleet may be assumed to be constant. We tested two schemes for T,^: a constant pattern of catchability over time for all fleets and catchability patterns for all fleets specific to four time periods (1=1971 Q4 to 1974 Q4, 2=1975 Ql to 1979 Q4, 3=1980 Ql to 1984 Q4, and 4=1985 Ql to 1992 Q4). Incorporating the notation for seasonal and multiyear effects, Equation 2 may then be rewritten as F,- ^s„tEjc l3j Quarterly effort statistics for fleets 1 and 3 were ob- tained from NMFS databases. Statistics for fleets 2 and 4 were obtained from Secretariat of the Pacific Community databases. An arbitrary, constant level of effort was assumed for fleet 5. Effort data were compiled for the period from 1971 Q4 to 1992 Q4 to correspond to the period of the tagging experiment. Changes in availability of the tagged population The model as described thus far implies that the availability of the tagged population to the various fisheries remains constant for the duration of the experiment. There are two reasons why this assump- tion may not be satisfied, necessitating some adjust- ment to the model. First, the spatial distribution of the tagged population of any release set in relation to the spatial distribution of fishing effort by the re- capture fleets was not constant over time. In par- ticular, because most releases occurred in the east- ern Pacific, the tagged population would have been initially more available to the U.S. baitboat and troll fleets and less available to the Japanese baitboat and longline fleets, which have a more westerly distribu- tion. Over time, as the tagged population dispersed, the effects of such differential spatial availability would be expected to dissipate. Second, the availabil- ity of the tagged population to the recapture fleets would also be affected by the size distribution of the tagged population. Tagged fish at release were gen- erally of a size similar to sizes of fish captured by the surface fisheries but were considerably smaller than those captured by the longline fishery. There- fore, the availability of the tagged population at re- lease would not be limited initially by size for the surface fisheries but would be reduced for the longline fishery. As the tagged fish grew, their availability to the longline fishery would become less restricted by their size, but their availability to the surface fish- eries might be reduced as they "grew out" of the size range typically exploited by those fisheries. Because our model has neither spatial nor size structure, the effects of size and spatial distributions cannot be explicitly incorporated. However, we developed an approximate means to allow for such effects in ag- 426 Fishery Bulletin 97(3), 1999 Table 2 North Pacific albacore tag- release and tag return data. summarized by release group. Releases Returns Date With recapture date Recapture date U.S. Japan Group (qtr/yr) Number baitboat baitboat Troll Longline Other unavailable Total 1 4/71 887 13 6 1 0 7 6 33 2 2/72 1110 42 10 2 3 11 15 83 3 3/72 976 48 3 4 2 27 14 98 4 2/73 1611 50 19 10 2 4 22 107 5 3/73 182 1 2 0 0 1 1 5 6 4/73 12 0 0 0 0 0 0 0 7 2/74 1379 39 23 5 2 3 12 84 8 3/74 1006 31 36 2 4 2 8 83 9 4/74 101 1 2 1 0 3 1 8 10 2/75 406 1 6 0 0 5 1 13 11 3/75 902 39 10 9 2 23 16 99 12 4/75 41 2 0 1 0 1 1 5 13 2/76 449 7 1 1 0 5 3 17 14 3/76 1115 32 7 6 3 11 14 73 15 4/76 17 0 0 0 0 1 0 1 16 1/77 1 0 0 0 0 0 0 0 17 2/77 63 1 2 0 1 0 1 5 18 3/77 1725 22 18 20 5 35 13 113 19 4/77 273 3 0 2 3 3 2 13 20 2/78 236 0 11 1 3 2 1 18 21 3/78 2196 21 29 56 15 27 16 164 22 4/78 290 2 1 1 2 3 1 10 23 2/79 272 0 8 0 2 0 1 11 24 3/79 1451 8 21 11 7 9 8 64 25 2/80 199 0 0 0 0 0 1 1 26 3/80 897 1 18 0 1 8 4 32 27 4/80 19 0 0 0 0 1 0 1 28 1/81 56 0 0 1 0 0 0 1 29 3/81 189 0 1 1 1 3 1 7 30 1/82 13 0 0 1 0 0 0 1 31 2/82 173 0 1 0 0 2 0 3 32 3/82 991 0 22 0 2 7 1 32 33 4/82 23 0 0 0 0 0 0 0 34 2/83 463 0 9 0 5 3 0 17 35 3/83 584 1 15 2 1 3 1 23 36 4/83 14 0 0 0 0 0 0 0 37 2/84 558 0 8 0 2 1 0 11 38 3/84 382 0 4 0 2 3 1 10 39 4/84 8 0 0 0 0 0 0 0 40 2/85 452 0 3 0 0 3 0 6 41 3/85 331 0 1 0 0 1 0 2 42 4/85 2 0 0 0 0 0 0 0 43 2/86 502 0 3 0 1 1 1 6 44 3/86 619 0 1 0 0 3 3 7 45 2/87 63 0 0 0 0 0 0 0 46 3/87 94 0 0 0 0 0 0 0 47 3/88 6 0 0 0 0 0 0 0 48 3/89 110 0 0 0 1 0 1 2 Total 23,449 365 301 138 72 222 171 1269 Bertignac et al.; Estimates of exploitation rates for Thunnus ala/unga from tagging data 427 gregate in order to reduce the potential bias in the other parameter estimates of interest. We defined a fleet-specific availability coefficient, 0,f, where the index t,j refers to the number of time periods at lib- erty of release group i in time period^. Equation (3) then becomes ^■Jf = l^'./'is.T, E.. (4) (p, f is therefore a proxy variable related to time at liberty that we used to correct, in an approximate way, for the effects of changing availability of the tagged population over time. However, 0, ^ need not be estimated for all t,j. Rather, ranges of ^^ may be speci- fied for which the 0, f are considered to be constant at either 0 (completely unavailable) or 1 (fully available). A scheme for constraining 0, f was devised from preliminary inspection of the data, and some knowl- edge of albacore dispersion (Laurs and Lynn, 1977), growth ( Laurs and Wetherall, 1981 ), and gear selec- tivity (Bartoo and Holts, 1993) in the North Pacific (Table 3). The scheme provided three time periods for spatial mixing of tagged albacore with respect to the surface fishery fleets and the "other" fleet (which has characteristics similar to the surface fishery fleets), and eight time periods to allow recruitment of tagged albacore to the longline fishery. During these periods, 0, ^ was estimated from the data. We assumed that tagged albacore remain fully available to the longline fishery during the constrained period for (<), / =1 for ^,,>8). For the surface fishery and "other" fleets, this may not be a good assumption, because these fishing methods may not have fully selected albacore of larger size (Bartoo and Holts, 1993). Therefore, we assumed that albacore were fully available during periods 4-11 after release {4>tj = 1 for 3<^,,<12) but were completely unavailable to these fleets after 15 time periods (0, / =0 for ^,,>15). Table 3 Scheme used to define I'anation in tagged albacore popu- lation availability. See text for definition of symbols . Fleet t,j (quarters) 01,/ estimated 0, , = 1 0,,/ estimated 0 ',/=0 U.S. baitboat 1-3 4-11 12- -15 >15 Japanese baitboat 1-3 4-11 12- -15 >15 Troll 1-3 4-11 12- -15 >15 Longline 1-8 >8 - - — Other 1-3 4-11 12- -15 >15 To allow for a gradual decline in availability during the intermediate period (ll<^y<16), (p^^ was esti- mated from the data. Parameter estimation We used a multinomial likelihood function to fit the various models to the tagging data. A derivation of the likelihood function as applied to tagging data is given in Kleiber and Hampton ( 1994). We minimized the negative log of this function to obtain the pa- rameter estimates, i.e. by minimizing (omitting terms dependent only on the data) -I (i?,-I/,.)ln R, X/,*ln (5) where the k subscript indicates an individual recap- ture stratum (combining recapture period, fleet, and time at liberty dimensions). Minimization was car- ried out with a quasi-Newton routine (Otter Re- search, 1991 ). The variance-covariance matrix of the estimated parameters was estimated from the in- verse of the Hessian matrix (Bard, 1974). The vari- ance of quantities that are functions of the estimated parameters (such as exploitation rates) were deter- mined by the delta method (Seber, 1973). Results Model fits Eight different model formulations, based on differ- ent combinations of constant or variable availabil- ity, constant or seasonal catchability, and presence or absence of multiyear effects on catchability, were fitted to the North Pacific albacore tagging data. The total numbers of estimated parameters and the maxi- mum log-likelihood function values for each fit are shown in Table 4. It is clear that the addition of vari- able availability, seasonal catchability, and multiyear effects on catchability all result in highly significant improvements in fit. The model incorporating all three effects, model 8, is suggested as the most ap- propriate of the tested models on the basis of likeli- hood-ratio tests (Kendall and Stuart, 1979). Model 6, which did not incorporate multiyear catchability ef- fects, was the next preferred model. Examples of plots of observed and predicted tag returns, aggregated over tag release groups, are shown in Figure 6 for model 2 and in Figure 7 for model 8. As expected, there are large discrepancies 428 Fishery Bulletin 97(3), 1999 Table 4 Maximum constant (- log-likelihood function values for alternative -) or seasonal (+) catchability, and absence (- model formulations con or presence (+) of multi prising constant (-) or variable ( + ) availability, -year effects on catchability. Model number Variable availability Seasonal catchability Multi-year effects on catchability Number of parameters Log likelihood 1 - — — 6 -9,511.2 2 - + - 19 -9,210.7 3 - - + 19 -9,386.1 4 - + + 50 -9,036.9 5 + - - 37 -9,089.9 6 + + - 47 -8,834.4 7 + - + 47 -8,973.3 8 + + + 79 -8,666.5 Table 5 Estimates of natural mortality rate M and 0, / obtained from fitting model 8 (see Table 4) to the North Pacific albacore tagging data. Underlined zeros indicate parameters set to zero because of zero tag return in those time periods. Para meter Recapt ure fleet U.S. baitboat Japan baitboat Estimate CV Troll Japan longline Estimate CV Other Estimat e CV Estimate CV Estimate CV Estimate CV M (1 year) 0.608 0.090 0,^ 0.176 0.207 0.035 0.510 0.408 0.252 0.000 — 0.774 0.174 ^2f 0.256 0.168 0.067 0.601 0.802 0.231 0.000 — 0.344 0.269 v 0.297 0.273 0.169 0.511 0.286 0.755 0.056 0.626 0.900 0.424 U 0.103 0.763 hf 0.000 — <^6f 0.414 0.470 fir 0.306 0.406 % 0.212 0.741 fiv 0.277 1.012 0.299 0.398 1.000 — 0.654 0.731 fu, 0.415 0.594 0.554 0.469 0.313 1.015 0.290 0.723 fur 0.486 0.470 0.237 1.018 0.000 — 0.679 0.600 fw 0.000 — 0.260 1.016 0.000 — 0.000 — between observed and predicted returns for model 2, particularly in the time periods soon after release. Major discrepancies were eradicated in the model 8 fit, which has greater flexibility in fitting the returns for these initial periods. Notice also that model 8 not only better predicts the returns for time periods in which predicted ♦ observed ■ti^aMMttMlMMiiA 1971 1973 1975 1977 1979 1981 1983 1985 1987 1989 1991 1993 Year Figure 6 Observed tag return numbers and tag returns predicted with model 2 (see Table 4 for description). higher than previous estimates obtained from fish- ery data by Suda (1966) and Fournier et al. (1998). Model 1-4 estimates of M (0. 496-0. 536/yr) are lower than those of models 5-8 (0. 576-0. 632/yr) because in the former, the models attempt to accommodate the low numbers of initial tag returns by depressing M, whereas in the latter, reduced availability for the initial time periods rather than reduced M is found to be a more likely solution. We noted that the corre- lation coefficients between the estimates of M and the 0, ^'s for the initial time periods were always negative and were in the range of -0.09 to -0.45. The coefficients of variation of the (^,y estimates were variable (0.16—1.0) and were particularly high for the surface fleets at time periods at liberty beyond three years when very few tagged albacore were recovered. For model 8, considerable differences in catch- ability estimates among time periods for all fleets were detected, although no consistent trends were evident (Fig. 8). The confidence intervals on these estimates were large in some cases. Estimates of the total exploitation rate (percent- age of the population harvested) in year y, calculated by 430 Fishery Bulletin 97(3), 1999 E 3 S CO S Z 100 n 80 60 40-1 20 0 U.S. baitboat - predicted Observed f*Mf>ftt»ft>tf ■■■!■■ Longline 0 12 16 20 24 28 32 36 40 - predicted observed 16 20 24 28 32 36 40 Japan baitboat -predicted ^^ observed ^g • y*»ifttifiiif nf ■■>■■ I 12 16 20 24 28 32 36 40 Other -predicted observed ^•^■^•••f Bf ■■■! ■■■f taf ■ I 16 20 24 28 32 36 40 Troll - predicted observed ■ ■yiiif"iif ■■■! mf af i 12 16 20 24 28 32 36 40 All fleets - predicted observed 12 16 20 24 28 32 36 40 Time at liberty (quarters) predicted « observed mMmmmmmmmA 1971 1973 1975 1977 1979 1981 1983 1985 1987 1989 1991 1993 Year Figure 7 Observed tag return numbers and tag returns predicted with model 8 ( see Table 4 for description ). Z^fZ^kti'^ti ' kf Z^/ Z^k^i'f'^v ' kf + M (6) l-expl-X/-! ,k^^%T„-^kf ■M X 100%, where /? is a quarter occurring in year j, are shown in Figure 9 (model 8, /3=0.9). Estimated exploitation rate has been <10% for most of the time series; it declined between 1976 and 1988 and increased slightly afterwards. Effect of assumed tag reporting rate Although the tag-reporting rate is thought to have been high for North Pacific albacore (Kleiber and Baker, 1987), this contention is not supported by independent data. We therefore examined the sensi- tivity of the results to different values of j3 (assumed to apply equally to each fleet) for the model 8 fit. The es- timates of M were directly related to p; conversely, av- erage exploitation rate was inversely related to /J (Fig. 10). In both cases, the sensitivity was slight for /3 > 0.6. If the tag-reporting rate was above this level for the main fleets, the results of our analysis should have been robust to small departures from the assumed value. Bertignac et al.: Estimates of exploitation rates for Thunnus alalunga from tagging data 431 In the absence of other informa- tion, the tag-reporting rate was as- sumed to be constant over time. It is possible that reporting could have varied over time, owing to variation in the effectiveness of tag recovery procedures or cooperation of fishing fleets. Although it is not possible to test conclusively such hypotheses with the available data, we might note that the time-related variabil- ity in catchability estimated with model 8 (Fig. 8) could equally have been interpreted as variation in re- porting rates, because catchability and reporting rate are highly cor- related in these models. Discussion U.S. baitboat Japan baitboat ■T— I J"l"l I I ' I — I-T — I I I I I I I 1234 1234 1234 Troll Longline Other Figure 8 Estimated relative catchability (seasonally averaged and normalized within fleet categories! by multiyear time period, with 95'* confidence intervals. 1=1971 Q4 to 1974 Q4, 2=1975 Ql to 1979 Q4, 3=1980 Ql to 1984 Q4, 4=1985 Ql to 1992 Q4. 12 e 10- 3 The use of a tagged sample of the population to infer characteristics of the population in general is a common technique in fisheries stock assessment. For such inferences to be valid, several assumptions, dis- cussed in detail by Seber (1973), need to hold. For North Pacific al- bacore, the assumption of equal probabilities of capture of tagged and untagged fish is likely to be critical. As noted earlier, both the spatial distribution of the tagged population in relation to recapture effort, and the size distribution of the tagged fish in relation to the size-se- lective characteristics of the different fishing gears are likely to result in violation of this assumption. We have developed a procedure to correct for these deficiencies in an approximate way. The procedure uses time-at- liberty information and knowledge of appropriate time lags to provide some correction for nonuniform avail- ability of the tagged population to the various fishing fleets that is due to spatial and size effects. A more elegant approach would be to develop a model that ex- plicitly deals with the spatial and size structure of the tagged population and with the spatial distribution and size selectivity of fishing effort. Spatially disaggregated tag models are now available (e.g. Kleiber and Hamp- ton, 1994; Sibertetal., in press). An extension of such a model to include size or age structure would pro- vide an improved method for analyzing the North Pacific albacore data. 14 T ii n T • - , T '•Ui 1972 1974 1976 1978 1980 1982 Year 1984 1986 1988 1990 1992 Figure 9 Time series and 95'* confidence intervals of estimated annual exploitation rates of North Pacific albacore obtained for model 8 (see Table 4 for description). The natural mortality rate parameter estimated from tagging data in this study is higher than some previous estimates for albacore. We have shown that M will be overestimated if the tag-reporting rate is overestimated. Although tag reporting for the major fleets is believed to have been high, it is possible that unreported tag recaptures by the drift gillnet fleet, which reported relatively few recaptures, have re- sulted in a positive bias in our estimate of M. If the exploitation of tagged albacore is similar to exploitation of the North Pacific population in gen- eral, our results suggest that aggregate exploitation rates declined from the mid-1970s to 1988, after which some increase occurred. The results of our preferred model (model 8) further suggest that an- 432 Fishery Bulletin 97(3), 1999 40. 30 - .2 20 - a. 10 - 0.0 nual total exploitation rates have been less than lO'^ during most of the years of the experiment. However, there is a strong indi- cation that exploitation of tagged albacore by the troll fleet may not be representative of exploitation patterns generally by this fleet. The total number of tag returns from the troll fleet was 148 (Table 1). The U.S. baitboat fleet, which fished in an area similar to that fished by the troll fleet during the 1970s and 1980s, reported 403 tag recaptures. However, during the period of the tagging experi- ment, troll catches were some eight times baitboat catches ( Liu and Bartoo, 1995). There is no ready explanation for this incon- sistency. One possibility is that differential reporting rates ex- isted between the fleets. How- ever, given the earlier assertion that tag-reporting rates were high in general, differential tag reporting is unlikely to be the cause of such a discrepancy. An- other possibility is that albacore become "hook shy" after they have been captured by trolling and are thereafter less likely to respond to trolled lures. Because the majority of albacore tag re- leases were troll-caught, such be- havior would render the tagged population less available to the troll fleet than the albacore popu- lation in general. If this was in- deed the reason for the paucity of troll-caught returns, relative availability of tagged albacore to the troll fleet would need to be of the order of 3.5% (compared with 100% for the U.S. baitboat fleet) to reconcile the tag-return and catch data for these fleets. Such reduced availability would result in significant underestimation of exploitation rates if not explicitly accounted for in the analysis. If we set (/)^^=: 0.035 for 32 individuals in at least 8 replicates) were analyzed with paired /-tests (P<0.05). Except for the weights and numbers of prawns and small-toothed flounder, all variables were analyzed by using one-tailed tests of the hy- pothesis that the square-mesh codends caught less than the control. Although the composite square- mesh codends might have been expected to retain fewer prawns than the conventional diamond-mesh control codend (due to the larger square-shaped mesh), Broadhurst and Kennelly ( 1996; 1997) showed that trawls fitted with composite square-mesh pan- els caught more prawns than conventional trawls (possibly owing to a reduction in drag and a corre- sponding increase in swept area). Therefore, the weight and number of prawns were analyzed by us- ing two-tailed tests. Catches of small-toothed floun- der, a species that cannot pass through the square meshes, were also compared by using two-tailed tests. With the exception of data for catches of prawns, where analyses provided similar results for weights and numbers of variables, only data concerning num- bers were included in Figure 4 to conserve space. Size frequencies of commercially or recreationally (or both) important fish were plotted and compared with two-sample Kolmogorov-Smimov tests (P=0.05), where there were sufficient data («>25 in each codend, pooled across all tows). Analysis of size-selectivity of prawns The relatively large size of mesh used in the control (45-mm diamond mesh), compared with the compos- ite square-mesh codends (52-mm mesh on the bar), meant that their selectivities overlapped. Insufficient numbers of smaller prawns were retained among individual hauls to enable analyses of between-haul variation (see Fryer, 1991). To provide sufficient data for analyses, size frequencies of prawns from each codend were combined across all tows. An estimated split model (Millar and Walsh, 1992) was used to fit logistic selection curves to these data by maximum likelihood method (Pope et al., 1975) by using the program "CC Selectivity" (Wileman et al., 1996). Logistic curve parameters, their standard errors, and 959^ confidence limits were calculated for each codend. For each logistic curve, model deviance val- ues were determined for a goodness-of-fit hypothesis (i.e. to test H^: that the curves were logistic). Size categories of commercially graded prawns from each codend were plotted and analyzed by us- ing two-sample Kolmogorov-Smirnov tests (P=0.05 ). In these analyses, the numbers of prawns pooled across all samples from each codend were used for the degrees of freedom. Results Analysis of catch data Compared with the control codend, both the compos- ite-square-2 and composite-square-3 codends signifi- Broadhurst et aL: Use and success of square-mesh codends in reducing bycatch and improving size-selectivity of prawns 439 ^ 2000 f 1500 = 1000 c (0 16/20 per pound). Discussion The work described in this paper illustrated the ef- fectiveness of a strategically located panel of square mesh in codends for reducing catches of juvenile fish (see also Robertson and Stewart, 1988; Fonteyne and M'Rabet, 1992; Briggs, 1992; Broadhurst and Kennelly, 1996, 1997) and quantified, for the first time in commercial penaeid prawn-trawls, the util- ity of large panels of square mesh for improving the size-selectivity of the targeted prawns. The composite square-mesh codends were equally effective in excluding large quantities of small indi- viduals of fusiform species, including sand trevally, red mullet, school whiting, and southern sand flat- head and there was no significant reduction in weights of prawns or total retained bycatch (Fig. 4; Table 1). Although the square mesh used through- out the main sections of both codends may have been large enough to permit some smaller fish (e.g. red mullet) to escape, the differences in the sizes of many fish retained between the composite square-mesh codends and the control (Fig. 5) indicated that to achieve the significant and quite large reductions in catch, most fish probably escaped through the larger 85-mm mesh panel located in the tops of the poste- rior sections of the codends. The escape of large numbers of these individuals through the 85-mm square-mesh panel in both codends may be attributed to the influences of water Broadhurst et al.: Use and success of square-mesh codends in reducing bycatch and improving size-selectivity of prawns 441 100 80 60 40 20 0 80n 60 40 20- - -nJlilll 100 H Composite square 2 n := 413 80 □ Control n = 630 60 40 20 • ■ ■ 0 rM»"i- B H Composite square 2 n = 53 □ Control n - 421 n-n-n-.-. 80 60 40 20 0 H Composite square 2 n = 298 80 Q Control n = 400 gQ ^ ■ Composite square 3 n = 355 n Control n = 476 tlll»i^ -^ ~ ^ ■ Composite square 3 n = 9 □ Control n = 422 B Composite square 3 n = 209 n Control n = 241 Figure 5 Size-frequency distributions of the composite square-mesh codends and control codend for lAi sand trevally [Pseudocaranx wrighti). (B) red mullet iUpeneichthys porosus), (C) Degen's leatherjacket iTIiamnaconus degeni). iDi school whiting iSillago bassensis). and (E) southern sand flathead iPlatycephaliis hassensisl flow anterior to the catch. The location of the panel (approx. 1.1 m from the end of the codend. Fig. 3) was based on a similar design (termed the compos- ite square-mesh panel) that is currently used com- mercially in oceanic prawn trawls in NSW where it has been shown to be very effective in reducing the bycatch of large numbers of small fish (Broadhurst and Kennelly, 1996, 1997). In previous experiments, Broadhurst and Kennelly (1996) and Broadhurst et al. (1999) determined that at this position, there is some displacement of water forwards owing to the twine area and build-up of catch in the posterior sec- tion of the codend. Because small fish are probably using anaerobic muscle power to maintain position in the moving trawl and are quite fatigued when they enter the codend, this displaced water may be suffi- cient 1) to assist them to swim forwards and out through the larger meshes in the panel; and (or) 2) to enable them to reduce their tail-beat frequencies and maintain their position in the vicinity of the larger mesh for a longer period, increasing their chances of randomly escaping. The extent to which such a flow facilitated the escape offish in the present study probably depended on their relative size or physiology (or both), because these factors largely influence swimming speed and endurance (Beamish, 1978). For example, the bycatch of fast-swimming species, such as sand trevally, and relatively large individuals of school whiting and southern flathead was greatly reduced in the two modified designs. However, although statistically significant, there was only a SO'^ reduction in the numbers of the relatively small Degen's leather jackets in the composite-square- 2 codend and a lower nonsignificant 24% reduction in the composite-square-3 codend (Fig. 41; Table 1). A possible explanation for the significant reduc- tion of Degen's leather jacket from the composite- square-2 codend may be the configuration of mesh 442 Fishery Bulletin 97(3), 1999 D A0-, 30 20- 10- 0 n. II II -II -II -I ■ Compostle square 2 n = 10 □ Control n = 212 10 15 20 25 40 30- 20- 10- 0 H Composite square 3 n = 7 ^ ContfOln= 194 10 n n IIJI II II ll^l II II ll-n n- 15 20 25 20 1 15 - 10 5-1 0 20 15- 10 - 5 - 0 -MJ 10 E M ■ Composite square 2 n = 19 □ Control n= 121 ■ jIIbJ.. H Composite square 3 n = 22 □ Control n= 167 ijilui III II n nt 15 20 25 Length (cm) 30 35 40 Figure 5 (continued) used in the posterior section of the codend and the effects that this had on distribution of catch and water flow. Because the posterior diamond-mesh sec- tion in this codend was 10 meshes in length, 100 meshes in circumference, and attached to a square- mesh section 62 bars in circumference (hanging ra- tio of 0.36), as the catch increased it would have spread laterally, increasing its surface area that was incidental to the flow and, therefore, increasing the displacement of water forwards (see Broadhurst and Kennelly, 1996, 1999). The composite-square-3 codend, however, was designed to increase size-se- lectivity of prawns and was tapered to a circumfer- ence of 58 bars attached to a posterior diamond-mesh section that was 2 meshes in length and only 58 meshes in circumference (hanging ratio of 0.57). Com- bined with lastridge ropes, this configuration would have forced a much smaller codend diameter than in the composite-square-2 codend, restricting any lat- eral distribution of catch. Any consequent reduction in surface area that was incidental to the flow and displacement of water forwards in this codend (which would limit any assistance to swimming fish) may account for the nonsignificant reduction of Degen's leather jackets and the poorer exclusion of small in- dividuals comprising discarded noncommercial bycatch (Fig. 4D). It should be noted, however, that these results were based on only 11 tows and that further experiments may be required to provide ad- ditional evidence to either support or refute the hy- pothesis discussed above. Although the mesh configuration and water flow in the composite-square-2 codend may have contrib- uted to the escape of slightly more fish, in terms of optimizing sizes of prawns retained, the composite- square-3 codend appeared to be a better design. Al- though the 95% confidence limits about the selectiv- ity parameters indicated no significant differences between the two designs (Fig. 6; Table 3), there was some evidence to suggest that the composite-square- 3 codend did appear to select slightly more commer- cial-size prawns. For example, compared with the control, this codend caught significantly fewer prawns by number (difference between means of Broadhurst et aL: Use and success of square-mesh codends in reducing bycatch and improving size-selectivity of prawns 443 Table 2 Length-frequency distributions (pooled across all tows) for prawns captured in the composite square-mesh and control codends, the obser\'ed selectivity (ratio of catches), and the fitted logit values. Carapace length in mm; CS2 = composite-square-2 codend; CSS = composite-square-3 codend. Carapace length CS2 codend Control codend Observed selectivity Logit value CS3 codend Control codend Observed selectivity Logit value 15 0 0 0.00 0.019 0 0 0.00 0.002 16 0 2 0.00 0.025 0 0 0.00 0.003 17 0 0 0.00 0.032 0 0 0.00 0.005 18 0 0 0.00 0.041 0 2 0.00 0.007 19 0 3 0.00 0.052 0 2 0.00 0.009 20 0 3 0.00 0.066 0 2 0.00 0.013 21 0 5 0.00 0.084 0 4 0.00 0.019 22 1 10 0.10 0.106 0 11 0.00 0.026 23 2 7 0.28 0.133 0 7 0.00 0.036 24 0 4 0.00 0.165 0 10 0.00 0.0.50 25 1 4 0.25 0.204 2 5 0.40 0.068 26 1 6 0.17 0.249 1 8 0.12 0.093 27 4 10 0.40 0.301 0 0 0.00 0.126 28 6 8 0.75 0.357 1 7 0.14 0.169 29 5 10 0.50 0.418 1 2 0.50 0.222 30 10 13 0.77 0.482 4 16 0.25 0.285 31 18 27 0.67 0.546 8 25 0.32 0.359 32 34 42 0.81 0.609 24 27 0.89 0.440 33 43 56 0.77 0.668 48 48 0.10 0.525 34 60 77 0.78 0.723 61 74 0.82 0.608 35 87 96 0.91 0.771 76 112 0.68 0.685 36 116 125 0.93 0.813 105 147 0.71 0.753 37 109 98 1.11 0.849 118 111 1.06 0.811 38 81 77 1.05 0.879 92 62 1.48 0.857 39 60 57 1.05 0.904 63 51 1.23 0.894 40 58 49 1.18 0.924 66 44 1.50 0.922 41 48 48 1.00 0.940 62 49 1.26 0.943 42 53 57 0.93 0.953 63 56 1.12 0.959 43 56 55 1.02 0.964 76 50 1.52 0.970 44 44 39 1.13 0.972 57 46 1.24 0.979 45 31 33 0.94 0.978 32 41 0.78 0.985 46 26 23 1.13 0.983 33 30 1.10 0.989 47 33 18 1.83 0.987 34 18 1.89 0.992 48 33 21 1.57 0.989 29 20 1.45 0.994 49 15 25 0.60 0.992 14 16 0.87 0.996 50 17 10 1.70 0.994 17 9 1.89 0.997 51 13 7 1.86 0.995 12 10 1.20 0.998 52 8 3 2.67 0.996 7 7 1.00 0.999 53 9 2 4.50 0.997 1 5 0.20 0.999 54 9 2 4.50 0.998 2 1 2.00 0.999 55 2 1 2.00 0.998 5 4 1.25 0.999 56 2 2 1.00 0.999 5 3 1.66 0.999 57 4 7 0.57 0.999 2 3 0.67 0.999 58 1 1 1.00 0.999 1 0 0.00 0.999 59 2 0 0.00 0.999 0 1 0.00 0.999 60 0 0 0.00 0.999 1 0 0.00 0.999 61 0 0 0.00 0.999 1 1 1.00 0.999 444 Fishery Bulletin 97(3), 1999 Table 3 Computed selectivity parameters for prawns (carapace length in mm) from the two composite square-mesh codends and deviance values for logistic curve goodness-of-fit. Standard errors are given in parentheses, a, b = logistic parameters (Pope et al.. 1975). P = split proportion from estimated split model (Millar and Walsh, 1992). Composite-square-2 codend Composite-square-3 codend b P 25'7f retention (L^j) 50% retention (Lj^) 75% retention (L,^) Selection range (SR) Deviance df P-value -7.803 0.257 0.544 26.02(1.19) 30.28(1.19) 34.56(1.76) 8.53(0.6) 32.48 39 0.760 -11.079 0.338 0.562 95% confidence limits 95% confidence limits 20.14-30.94 29.46(0.76) 25.42-33.07 28.13-33.38 32.71(0.82) 31.29-34.59 29.73-42.20 35.95(1.13) 32.38-40.91 4.89-12.16 6.49(0.5) 46.54 40 0.221 4.21-8.76 12.1%) than did the composite-square-2 codend (difference between means of 8.1%), and there was no significant reduction in weights (Fig. 4; Table 1). These sHght differences in size-selectivity of prawns may be explained in terms of the be- havior of prawns in the codend and the differ- ent configurations of mesh discussed above. Previous studies have shown that the response of prawns to stimuli from the trawl is mini- mal and after initial contact with the leading edge of the footrope, the prawns are quickly forced into the rear of the codend (Lochhead, 1961;Watson, 1976). Their escape at this point is primarily determined by their probability of randomly encountering openings between meshes that are large enough to pass through. In the composite-square-2 codend, the hang- ing ratio between the posterior diamond and square-mesh sections would have restricted the fractional diamond-mesh openings, limit- ing the opportunity for escape. Further, as the catch increased and spread laterally, this would have widened the diameter of the posterior section, pro- viding prawns with less opportunity of encountering any open meshes in the sides of the codend. In con- trast, we designed the composite-square-3 codend, so that it would assume a smaller diameter and cir- cumference during fishing, but with a greater area of square mesh, thereby providing more opportunity for prawns to randomly encounter the sides of the codend and the open square meshes. The results obtained in this study showed that composite square-mesh codends in penaeid prawn trawls can improve size-selectivity of the targeted 09 X^"""^^ 08 / / -O 07 a> c eg 06 C 05 1 04. / / ^ / Composite square 2 o Ol 03 02 ■ 01 / / 15 20 25 30 35 40 45 50 55 60 Carapace length (mm) Figure 6 Size-selectivity curves for king prawns iPenaeus latisulcatiis) from the composite square-mesh codends. prawns while reducing large quantities of unwanted fish. It is also apparent, that such modifications may improve the overall efficiency of the trawl in terms of increasing catches of prawns. For example, al- though the composite-square-2 and square-3 codends significantly reduced the numbers of prawns caught by 8.7% and 12.1%, compared with the control, the differences in weights of prawns were much smaller, to the point where they were nonsignificant (e.g. 1.8% and 3.7%, respectively). To achieve such a result, the two trawls containing composite square-mesh codends caught more prawns but retained mainly larger-size individuals. Broadhurst et al.: Use and success of square-mesh codends in reducing bycatch and improving size-selectivity of prawns 445 D 25 2.0 1.5 1.0 05 00 -05 -1.0 -1.5 -2 0 -2 5 25 20 1.5 1.0 0.5 OO -0 5- -1.0- -1.5 -2 0-1 -2 5 I .III I ■ 1 T' ■ lll'l B TTT| r M ML 15 20 25 30 35 40 45 50 55 60 Carapace length (mm) Figure 7 Plots of the deviance residuals from selection curves for the (A) composite-square-2 codend and (Bl composite-square-3 codend. One hypothesis to explain this result is that the large, open, square meshes in the composite square- mesh codends, combined with the overall size of the codends (approx. 4 m in length) allowed a faster re- lease of water from the body of the trawl than did the control. Such an increase in flow may have re- sulted in prawns quickly passing into the codend af- ter initial contact with the footrope, with less chance of randomly escaping over the headrope or out through the mouth of the trawl. In support of this, Walsh et al., (1992) provided evidence of similar flow-related ef- fects on the catches offish (American plaice, Hippo- glossoides platessoides) from trawls fitted with square-mesh codends. In the present study, however, any potential effects of increased flow did not appear to influence fish such as small-toothed flounder (the only species that could not pass through the square mesh), because there were no significant differences in the catches between the composite square-mesh and control codends ( Table 1 ). An alternative hj^joth- esis proposed by Broadhurst and Kennelly (1997) to describe a similar increase in catches of prawns from NSW oceanic prawn trawls containing codends with composite square-mesh panels is that by reducing the weight of bycatch and therefore the drag in the codend, trawls with the composite square-mesh achieved slightly more swept area than did controls, thereby covering more of the sea bed and capturing more prawns. Regardless of the underlying mechanisms, the in- crease in catches of target-size prawns, combined with the substantial reductions in the bycatch offish, 446 Fishery Bulletin 97(3), 1999 led to unanimous industry acceptance and adoption of the composite-square-3 codend within two weeks of the conclu- sion of the field work described in this paper. Fishermen have since reported that the codend has minimal distortion (due to the lastridge ropes), allows in- creased tow duration, and provides an improved quality of prawns. These re- sults illustrate the benefits that can be derived through liaison with industry and by incorporating their ideas into modifications to improve the selectivity of prawn trawls. Such voluntary adop- tion should ensure the continued devel- opment and refinement of designs, as part of normal commercial operations. Acknowledgments Funding for this work was provided by the Australian Fishing Industry Re- search and Development Corporation. We are grateful to Jim Raptis (A. Raptis & Sons PTY LTD ) and Yvonne Aston for the use of their vessel Jillian Sandra. Thanks are extended to Geoff Gordon, Bruce Jackson, Kim Redman, Sam Tudorvic, Micheal Aston, Bill Walsh, Les Lowe (Gulf Net Mending PTY LTD), Hec Kavenagh, and all Gulf St. Vincent prawn-trawlers for their advice, assis- tance, and support. We would also like to acknowledge the Norwegian Research Council for their contribution to the in- volvement of Roger Larsen in this work. 25 20- 15 10- I Composite square 2 n = 4846 □ Control n = 5448 ItL B 25- 20- 15 10 5- B Composite square 3 n = 4268 □ Control n = 4495 U/6 6/8 u/10 9/12 13/15 16/20 21/30 Commercial prawn grades by weight (no, per pound) Figure 8 Commercial size categories of king prawns iPenaeus latisulcatus) caught witli the (A) composite-square-2 and control codends and (Bi composite- square-3 and control codends (u=underl. Literature cited Alverson, D. L., M. H. Freeberg, S. A. Murawski, and J. G. Pope. 1994- A global assessment of fisheries bycatch and discards. FAO Fish. Tech. Paper 339, Rome, 233 p. Andrew, N. L., K. J. Graham, S. J. Kennelly, and M. K. Broadhurst. 1991. The effects of trawl configuration on the size and com- position of catches using benthic prawn trawls off the coast of New South Wales, Australia. ICES J. Mar. Sci. 48:201- 209. Andrew, N. L., and J. G. Pepperell. 1992. The by-catch of shrimp trawl fisheries. Oceanogr. Mar. Biol. Annu. Rev. 30; 527-565. Beamish, F. W. H. 1978. Swimming capacity, /fi W. S. Hoar and D. J. Randall (eds.), Fish physiology, vol. VII, Locomotion, p. 101- 187. Academic Press, New York. NY. Briggs, R. P. 1992. An assessment of nets with a square mesh panel as a whiting conservation tool in the Irish Sea Nephrops fishery. Fish. Res. 13:133-152. Broadhurst, M. K., and S. J. Kennelly. 1996. Effects of the circumference of codends and a new design of square-mesh panel in reducing unwanted by-catch in the New South Wales oceanic prawn-trawl fishery, Australia Fish. Res. 27:203-214. 1997. The composite square-mesh panel: a modification to codends for reducing unwanted bycatch and increasing catches of prawns throughout the New South Wales oce- anic prawn-trawl fishery. Fish. Bull. 95:653-664. Broadhurst, M. K., S. J. Kennelly, and S. Eayrs. 1999. Flow-related effects in prawn-trawl codends: poten- tial for increasing the escape of unwanted fish through square-mesh panels. Fish Bull 97(11:1-8. Fonteyne, R., and R. M'Rabet. 1992. Selectivity experiments on sole with diamond and Broadhurst et al.: Use and success of square-mesh codends in reducing bycatch and improving size-selectivity of prawns 447 square mesh codends in the Belgian coastal beam trawl fishery. Fish. Res. 13:221-233. Fryer, R. J. 1991. A model of between-haul variation in selection. ICES J. Mar. Sci. 48:281-290. Karlsen, L., and R. Larsen. 1989. Progress in the selective shrimp trawl development in Norway. In C. M. Campbell (ed.), Proceedings of the world symposium on fishing gear and fishing vessels, p. 30-38. Marine Institute, St Johns, Canada. Kennelly, S. J., and M. K. Broadhurst. 1995. Fishermen and scientists solving bycatch problems: Examples from Australia and possibilities for the north- eastern United States. In Solving bycatch: considerations for today and tomorrow, p 121-128. Alaska Sea Grant College Program Report 96-03, Univ. Alaska, Fairbanks, AK. Lochhead, J. H. 1961. Locomotion. In T. H. Waterman (ed.). The physiol- ogy of the Crustacea, 681 p Academic Press, New York, N\'. Millar, R. B., and S. J. Walsh. 1992. Analysis of trawl selectivity studies with an applica- tion to trouser trawls. Fish. Res. 13:205-220. Pope, J. A., A. R. Margetts, J. M. Hamley, and E. F. Akyuz. 1975. Manual methods for fish stock assessment. Part III: Selectivity of fishing gear FAO Fisheries Technical Re- port 41 (revision 1), 6.5 p. Renaud, M., G. Gitschlag, E. Kilma, A. Shan, D. Koi, and J. Nance. 1993. Loss of shrimp by turtle excluder devices (TEDs) in coastal waters of the L'nited Sates, North Carolina to Texas: March 1988-August 1990. Fish. Bull. 91:129-137. Robertson, J. H. B., and P. A. M. Stewart. 1988. A comparison of size selection of haddock and whit- ing by square and diamond mesh codends. J. Cons. Int. Explor Mer 44:148-161. Robins-Troeger, J. B., R. C. Buckworth, and M. C. L. Dredge. 1995. Development of a trawl efficiency device (TED) for Australian prawn fisheries. II. Field evaluations of the AusTED. Fish. Res. 22:107-117. Rogers, D. R., B. D. Rogers, J. A. de Silva, and V. L. Wright. 1997. Effectiveness of four industry-developed bycatch re- duction devices in Louisiana's inshore waters. Fish. Bull. 95:552-565. Suuronen, P., and R. B. Millar. 1992. Size selectivity of diamond and square mesh codends in pelagic herring trawls: Only small herring will notice the difference. Can. J. Fish. Aquat. Sci. 49:2104-2117. Thorsteinsson, G. 1992. The use of square mesh codends in the Icelandic shrimp (Pandalus borealis'i fishery. Fish. Res. 13:255- 266. Tokai, T., H. Ito, Y. Masaki, and T. Kitahara. 1990. Mesh selectivity curves of a shrimp beam trawl for southern rough shrimp Trachypenaeus curvirostris and mantis shrimp oratosquilla oratorio. Nippon Suisan Gakkaishi 56:1231-1237. Tokai, T., and H. Sakaji. 1993. Mesh selectivity of a shrimp beam trawl for tora vel- vet shrimp Metapenaeopsis acclivis, kishi velvet shrimp Metapenaeopsis dalei and smoothshell shrimp Para- penaeopsis tenella. Bull. Nansei Natl. Fish. Res. Inst. 26:21-30. Walsh, S.J., R. B. Millar, C. G. Cooper, and W. M. Hickey. 1992. Codend selection in American plaice: diamond ver- sus square mesh. Fish. Res. 13:235-254. Watson, J. W. 1976. Electrical shrimp trawl catch efficiency for Penaeus duorarum and Penaeus aztecus. Trans. Am. Soc. 105: 135-148. Watson, J. W., J. F, Mitchell, and A. K. Shan. 1986. Trawling efficiency device: A new concept for selec- tive shrimp trawling gear Mar Fish. Rev. 48(1): 1-9. Wileman, D. A., R. S. T. Ferro, R. Fonteyne, and R. B. Millar. 1996. Manual methods of measuring the selectivity of towed fishing gears. ICES, Copenhagen, 126 p. 448 Fishery Bulletin 97(3), 1999 Appendix The plots of deviance residuals (Fig. 7) for both se- lection curves provided in Figure 6 showed that most of the size classes of prawns above 45-mm carapace length had high deviances. By removing these data from analysis and also the frequencies for the 35- and 36-mm size classes from the composite-square-3 codend, the P-values for model goodness-of-fit in- creased from 0.760 to 0.999 for the composite-square-2 codend (df=25) and from 0.221 to 0.853 for the com- posite-square-3 codend (df=21) (Table 4). Although this alternative analysis does provide selectivity pa- rameters that are slightly different from those cal- culated in Table 3, the overall interpretation of the data remains the same. Table 4 Computed selectivity parameters for prawns (carapace length in mm) from the two composite square-mesh codends and deviance | values for logistic curve goodness-of-fit, excluding all entries above 45-mm carapace length for both codends and the 35- and 36-mm size classes from the composite -square-3 codend. Standard errors are given in parentheses, a.b = logistic parameters (Pope et al., | 1975). P = split proport on from estimated split model (Millar and Walsh, 1992). Composite-square-2 codend Composite- square-3 codend a -9.556 -13.58 b 0.330 0.424 P 0.513 95% confidence limits 0.573 95% confidence limits 2b'7c retention (Lj^) 25.59(0.60) 21.80-29.17 29.46(0.59) 26.33-32.38 50"^ retention (Ljg) 28.92(0.63) 27.73-30.42 32.05(0.66) 30.90-33.67 75% retention (L75) 32.24(0.92) 28.68-36.64 34.64(0.96) 31.77-38.67 Selection range (SR) 6.65(0.92) 4.84-8.45 5.18(0.81) 3.42-6.95 Deviance 7.91 14.75 df 25 21 P-value 0.999 0.853 449 Abstract.— A gross measure of repro- ductive condition (ovary weight ad- justed for body size and oocyte volume) is developed and evaluated as an alter- native to commonly used gonad indices, for classifying the maturity status of in- dividual ehu {Etelis carbunculus) and kalekale tPnstipomoides sieboldii), two species of eteline snappers ( Lutjanidae ) that contribute to the deep-slope hand- line fishery in Hawaii. Discriminant analysis and logistic regression, based on body length, ovary weight, and oo- cyte volume, were used to classify fish as either immature or mature. Dis- criminant analysis correctly classified the maturity of about 98% of both ehu and kalekale, with histological criteria as the standard for comparison. Logis- tic regression correctly classified ma- turity of 97% of the ehu and 100% of the kalekale. Misclassification errors increased by 3.75-5% (discriminant analysis) or 0-5% (logistic regression) if oocyte volume was excluded and only body length and ovary weight were used as predictors of maturity. For kalekale, estimates of lengths at which 50% are sexually mature 'L^yi were identical (29.0 ±1.8 [SE] cm fork length, FL; r-=0.92), when maturity was classified histologically or by logis- tic regression on body, ovary, and oo- cyte metrics. For ehu, L^^,, estimates were similar (r2=0.96l for maturity as- signments based on histology (27.9 ±2.4 cm FL) and on logistic regression us- ing gross metrics (27.8 ±2.3 cm FL). We conclude that gross morpho- metries can provide adequate proxies for histological evidence when catego- rizing sexual maturity in asynchro- nous, multiple-spawning fishes like eteline lutjanids. The potential benefits of using gross metrics for assessing sexual maturity in other serial spawn- ers are briefly discussed. Morphometric criteria for estimating sexual maturity in two snappers, Etelis carbunculus and Pnstipomoides sieboldii Edward E. DeMartini Boulderson B. Lau Honolulu Laboratory Southwest Fisheries Science Center National Marine Fisheries Service, NOAA 2570 Dole Street Honolulu, Hawaii 96822-2396 E-mail address (for E E DeMartini) Edward DeMartiniiginoaa gov Manuscript accepted 28 August 1998. Fish. Bull. 97:449-458 ( 1999). Body size at sexual maturity is of fundamental importance in fishery stock assessment (e.g. when calcu- lating a "spawning potential ratio" [SPR, Goodyear, 1993; Somerton and KobayashiM '• In fisheries ap- plications, average size at maturity is usually expressed as the body length at which 50 percent of fe- males are mature (Ljg), and is typi- cally estimated by fitting a model such as the logistic (Gunderson et al., 1980; O'Brien et al., 1993) to observed percentages of mature fish within several length classes. Esti- mating percentage mature by size class assumes that individuals can be accurately classified as either immature or mature. Various methods have been used to classify individual female fish as sexually mature, including gross inspection of ovaries, light micro- scopic examination and measure- ment of whole oocytes in fixed tis- sue samples, histological stains of intraovarian inclusions, and an ar- ray of ratios, commonly referred to as "gonad indices" or "gonadoso- matic indices" (GSIs) that standard- ize gonad mass to body size. These methods differ in precision, accu- racy, and processing time. Gross inspection of ovaries is quick but highly subjective and often cannot be used to distinguish between im- mature and spent (resting) mature fish (West, 1990). Light microscopy of whole oocytes can effectively identify the most advanced oocyte mode only if the dynamics of oocyte growth are known for the particu- lar species and may be inaccurate for ovaries that also contain oocytes in intermediate stages of vitellogen- esis (West, 1990). Histological stain- ing of tissue samples for detection of cytoplasmic inclusions (yolk, oil globule presence and number) in the most advanced oocytes provides the best evidence of maturation and is the most accurate method, but it is also the most time consuming and expensive (Delahunty and deVla- ming, 1980). Usually only a single method has been used in a given study, although some studies (e.g. Cayre and Laloe, 1986; Hay et al., 1987; Kjesbu, 1991; Olsen and Rulifson, 1992; Ramsay and Witthames, 1996) have at least attempted to confirm maturity clas- sifications with more than one tech- nique. Average size at maturity has usually (e.g. Higham and Nicholson, 1964; DeMartini and Fountain, 1981), but not always (Erickson et al., 1985a; Matsuyama et al., 1987; McQuinn, 1989; Hunter etal., 1992; ' Somerton, D. A., and D. R. Kobayashi. 1990. A measure of overfishing and its application on Hawaiian bottomfishes. Ho- nolulu Laboratory, Southwest Fish. Sci. Cent., Natl. Mar Fish. Serv., NOAA, Ho- nolulu, HI 96822-2396. Southwest Fish. Sci. Cent. Admin. Rep. H-90-10, 18 p. 450 Fishery Bulletin 97(3), 1999 Render et al., 1995), been identified subjectively if more than one method of classification was used, sometimes with inconsistent results (Trippel and Harvey, 1991). Recently, West (1990) identified mi- croscopic staging coupled with a gonad index as the most objective and quantitative method. Gonad indices should be used with caution as mea- sures of sexual maturity and require biologically re- alistic companion analyses (e.g. Koya et al., 1995). A fundamental flaw of ratios, such as gonad indices, is that they either lose or obscure meaningful informa- tion in the numerator and denominator of the ratio (Garcia-Berthou and Moreno-Amich. 1993). Gonad indices also are questionably applicable for asynchro- nous, multiple-spawning fishes (deVlaming et al., 1982). There are two major reasons for this. First, because multiple-spawners ripen and shed numer- ous batches of eggs, the ovary weight of an individual fish fluctuates cyclically and often greatly during the spawning season. Second, somatic weight often var- ies independently of ovary weight for individuals. Somatic "condition" (robustness) usually declines monotonically, and sometimes greatly, as somatic energy reserves become progressively depleted over the spawning season, whereas ovary weights cycle high and low as the individual continues to spawn (Delahunty and deVlaming. 1980; deVlaming et al., 1982). Variances in GSIs among mature fish are un- avoidably large (e.g. Macer. 1974) because small, ready-to-spawn fish sometimes have higher GSIs than larger fish that have just shed their eggs. Other potential complications include differences in ova- rian cycling frequencies (spawning intervals) be- tween small and large females within populations ( Hunter and Macewicz, 1985 ). Thus even refinements of the conventional GSI such as the "relative gonadal index" (RGI) of Erickson et al. (1985b), in which go- nadal investment is adjusted for size allometry, can be inappropriate for asynchronous, multiple-spawn- ers if the dynamics of somatic and gonadal condition are unknown. Statistical descriptors of sexual ma- turity that are more accurate and robust than GSIs need to be developed (Garcia-Berthou and Moreno- Amich, 1993; Winters and Wheeler, 1996; Jons and Miranda, 1997). Our study has several complementary objectives. First, we evaluate four gross metrics ( fish body length and weight, ovary weight, and oocyte size) for de- scribing stages of sexual maturation in two eteline lutjanids (ehu, Etelis carbunculus, and kalekale, Pristipomoides sieboldii) that contribute to the deep- slope bottom fishery in Hawaii. We examine these four characters because they are readily measured and less costly than the more accurate histological staining of ovarian tissue. We also independently assess stage of sexual maturity for each species based on histology of ovarian tissue samples from speci- mens collected during peak periods of spawning. In our study, we use discriminant analysis and logistic regression to further classify individual fish as ei- ther immature or mature on the basis of gross mor- phometric and histological criteria. In conclusion, we evaluate the results of gross classifications of matu- rity for ehu and kalekale by assessing how well L^g estimates based on statistical assignments of matu- rity with gross metrics agree with L^q estimates based on histological evidence of maturity. Materials and methods Fish collection and shipboard processing Two types offish collections were used: hook and line and bottom trawling. For exploitable-sized fish of both species, hydraulic handline gurdies with sizes 28 and 34 Izuo circle hooks ( baited with squid strips ) were used to fish near bottom at depths of 60-300 m, on insular slopes and submerged banks located be- tween 22-25=N lat. and 160-170°W long, in the Northwestern Hawaiian Islands (NWHI). Specimens were collected in September 1992; during June, July, August, and September 1993; in August 1996, and in July and August 1997 (Table 1). Most Une-fishing was conducted during daylight hours (0730-1930) aboard the NOAA ship Townsend Cromwell. Addi- tional small kalekale were sampled at 60-65 m off south Molokai, Main Hawaiian Islands, during Au- gust 1992 and September 1993. We also examined additional small ehu caught by bottom trawl at 88- 97 m near Kure Atoll (NWHI) in early June 1977 (Table 1). All specimens examined were collected during the protracted summer spawning period of eteline snappers in Hawaiian waters (Ralston, 1981; Everson, 1984; Everson et al., 1988). Ehu are thought to be serial spawners (Everson, 1984), and oocyte size distributions within ovaries suggest that Hawaiian populations of both species comprise asynchronous spawners.- Once captured, fish were placed immediately on ice. Within an hour of capture, fork length (FL) to the nearest 0.1 cm and ovary-free body weight (OFBW, in g) was estimated for each fish. Ovary weight ( OW) of each specimen that appeared female was estimated ±0.1 g (if <5g), ±1 g (if 5-100 g) and ±10 g (if >100 g). A tissue plug (1-50 g, de- - 1994. Honolulu Laboratory. Southwest Fish. Sci. Cent.. Natl. Mar Fish. Serv., NOAA, Honolulu, HI 96822-2396. Unpubl. data. DeMartini and Lau: Criteria for size at maturity in snappers 451 Table 1 Summary statistics for ehu ( Etelis carbuncul us) and kalekale Pnstipomoi des sieboldii) used in analyses Months were sampled | in the years 1992, 1993, 1996, and 1997. FL = fork length; OFBW = ovary- free body weight; OW = ovary weight. Species Month FL(mm) OFBW (g OW(g) n Median Min- -Max Median Min -Max Median Min- -Max Ehu Jun-Jul 368 91- -551 865 15- -2984 10 0.1- -136 37 Aug 356 203- -544 826 200- -3260 24 0.3- -350 80 Sep 370 270- -595 894 351- -3615 28 3.0- -200 64 Total 362 91- -595 850 15- -3615 23 0.1- -350 181 Kalekale Jun-Jul 340 334- -390 714 596- -1054 40 26-46 5 Aug 340 156- -438 626 51- -1400 20 0.1- -120 51 Sep 363 150- -412 846 51- -1247 17 0.1 -39 24 Total 348 150- -438 710 51- -1400 20 0.1- -120 80 pending on ovary size, and including the ovarian wall) was taken from the central one-third of either the right or left ovary (random choice) and placed in 107c formalin sea-water. Laboratory processing Most ovary specimens were processed within 1-2 months after collection. Diameters of 25 of the larg- est viable oocytes were then measured (random axis, 63x magnification) for each specimen with a dissect- ing microscope. The median of 25 diameters provides a cost-efficient estimator of average maximum oo- cyte size for ehu (Lau and DeMartini, 1994). For each preserved ovary specimen, a subsample of the ovary was dehydrated, imbedded in paraffin, and a minimum of three sections was cut (at 6 |im) and stained with Harris's hematoxylin, followed by eosin counterstain (Hunter and Macewicz, 1985). Slide sections were examined with a compound mi- croscope at 40-500x for the presence and relative quantity of eosinophilic yolk ( unyolked oocytes: class 1 of Murphy and Taylor, 1990; partly yolked: classes 2-3; yolked: classes >4). If oocytes were yolked, we further noted the presence of postovulatory follicles (POF: class 6), hydrating or hydrated oocytes (HYD: class 7), and oocytes undergoing a or /3 atresia (class 8: Murphy and Taylor, 1990). Individuals were des- ignated as "immature" if the most advanced oocytes were unyolked or partly yolked without substantial atresia, "ripening mature" if fully yolked but lacking POF or HYD, "resting mature" if a majority of the yolked oocytes were atretic, or "ripe mature" if POFs or HYDs were present. Fish that might in the future have resorbed yolked oocytes without spawning were necessarily classified as mature. Statistical analyses We used oocyte volume (OV) to characterize egg size because a volumetric measure was more directly pro- portional to OW than was a linear measure such as oocyte diameter. OV was calculated as the volume of a sphere, 4/3 m-^. where r was one-half of the median diameter of the largest oocytes (Lau and DeMartini, 1994). The sampling distributions of dependent and regressor variables were first examined by using raw and log-transformed data. Log-transformed data were least skewed; logarithms (base 10) were there- fore used in all subsequent analyses. We used analysis of covariance (ANCOVA) to de- termine if OW was suitable for quantifying matura- tion state of individual fish after adjusting OW for body size, egg size, or both measures. For each spe- cies, the analysis evaluated fish of three ovarian de- velopmental stages (unripe immature, ripening or resting mature, and ripe mature), identified histo- logically. We were thereby able to evaluate the po- tentially confounding effects of oocyte development and body size on ovary weight ( Erickson et al., 1985b). We further used two classification techniques (dis- criminant analysis and logistic regression) to distinguish between mature and immature fish. These two techniques were used because we recognized the general need, in studies such as this, to classify the maturity status of individual fish 1 ) based on both gross morphometries and histological criteria and 2 ) with gross metrics alone, without the added benefit of histological evidence. At a minimum, ovary tissues of some of the fish examined in any study need to be examined histologically as a check on the accuracy of gross classification. Both logistic regression and discriminant analy- sis can be used to classify dichotomous states such 452 Fishery Bulletin 97(3), 1999 as maturity. The latter is often (Efron, 1975), but not always (Prager and Fabrizio, 1990), the more effec- tive technique. However, it is less robust to viola- tions of the assumption of identical covariance ma- trices. Multivariate normality is also assumed if lin- ear discriminant functions are fitted parametrically (Press and Wilson, 1978), an assumption we avoided in this study (see below). For combined gross-histological classifications of maturity, we used predictive discriminant analysis (Huberty, 1994) to distinguish between mature (ripe, ripening or resting) and immature fish. Nonparamet- ric discriminant functions were derived by using the uniform-kernel method with Euclidean distances (Huberty, 1994). We used logistic regression (Hosmer and Lemeshow, 1989) to evaluate whether maturity classifications based on histology could be predicted by the relationships among body size, OW, and OV. Logistic regression coefficients were estimated by maximum likelihood methods. Using each technique, we evaluated the full (three-variable) model first, then a two-variable model without OV. Because sample sizes were small for each species, all sample fish of each species were evaluated histologically. Once all individuals had been assigned to matu- rity classes based on histological and gross morpho- metric criteria, we assigned specimens to 2-cm length classes and calculated the percentage mature for each length class using each set of criteria. Percent matu- rity (0-100%) was then related to length class by using nonlinear regression weighted by the square root of the numbers offish in each length class. The logistic model, P, = 100/(l+e-'°"''^^'), where P = percentage mature in each length class; a and b = fitted parameters; and L^Q - (-a/6), was fitted to the data by using maximum likelihood. Data were analyzed using STATISTIX ( v 4. 1 ; Ana- lytical Software, 1994: logistic regression) and SAS (V 6.03; SAS Institute, Inc., 1988: PROC REG, GLM, andDISCRIM). in Table 2 and used in all subsequent analyses. OFBW was significantly related to FL for each species (ehu: logOFSW=-1.720 + 2.9711ogFL, r-'=0.979, n = 172, P<0.001; kalekale: logOFBW=-1.785 + 3.0101ogPL, r-'=0.966, n=75, P<0.001). Effect of ovarian stage on relation of ovary welgfit to body length The OW-to-body length relation marginally differed with ovarian developmental stage for ehu and kalekale (FL x stage interaction in ANCOVA; P., j,5=3.19 and ^2,74=2-69, P=0.04 and 0.06, respec- tively; reject Hq. slopes equal). Thus, ovary weights could not be adjusted by body lengths and subse- quently used to distinguish among maturation states for either species. Body length was a poor predictor of ovary weight for both ehu and kalekale (r-=0.58 and 0.60, respectively; Table 2). Effect of egg size FL and OV together, however, accounted for 91% (ehu) and 82.5% (kalekale) of the variation in OW (Table 2). Colinearity between FL and OV (or OFBW and OV) was unimportant. For example, FL ex- plained only 12% of the variation in OV for ehu. Histological evidence of maturity Most specimens of ehu and kalekale were readily classified as either immature or mature according to histological criteria. Kalekale included fish that were immature, ripening mature, and ripe mature (none were resting mature). Unlike kalekale, a mi- nority of the large (>40 cm FL) ehu were resting mature, with ovaries undergoing substantial atre- sia. The ovaries of kalekale usually contained HYD oocytes if POFs were present (19 of 21 cases). Ehu included specimens with HYD oocytes (7 cases) or POFs ( 98 cases ), but POFs were not observed if HYDs were present (0 of 7 cases). Ehu and kalekale clearly differed in the co-occurrence of HYD oocytes and POFs (Fisher exact test; P<0.001), suggesting that the average spawning intervals of individual females probably differ between these two species. Results Body size interrelations After log-transformation, the influences of body length (FL) and ovary-free body weight (OFBW) on ovary weight (OW) were virtually identical; hence, FL (the easier variable to measure) only is reported Statistical classifications of maturity For both species, maturity classifications based on discriminant analysis of the three gross metrics (body length, ovary weight, and oocyte volume) generally agreed (=98% correct) with those where histological criteria were used (Table 3). Maturity classes deter- mined by histology also were accurately predicted DeMartini and Lau: Criteria for size at maturity in snappers 453 Table 2 Multiple regression models describing the interrelations of ovary weight (OW), fork length (FL). and oocyte volume (OV) for two species of eteline lutjanids in Hawaii iehu. Etelis carbunculus: ka\eka\e, Pristiponioides sieboldii). Mean square errors (MSE)and standard errors of estimates (SE 1 are noted, with coefficients and standard errors (SE) of model parameters expressed as base 10 logarithms. All regressions are significant, as are the intercepts (all P<0.001). Species Model R2 n MSE Coefficient SE Intercept Y = flX^+X^+X^X.^) b. b. il2 SEm SE,, SEbl2 Ehu OW = f\FL\ 0.579 172 0.102 4.541 — — 0.297 -5.839 OW = f(FL + OV) 0.911 172 0.022 3.284 0.457 — 0.146 0.018 — -5.508 OW = flFL + OV + FLX OV) 0.917 172 0.020 1.077 -.501 0.627 0.646 0.274 0.179 -2.149 Kalekale OW = flFL) 0.596 75 0.111 7.007 — — 0.676 — — -9.579 OW = flFL + OV) 0.825 75 0.049 3.877 0.376 — 0.551 0.039 — -6.297 OW = flFL + OV + FL - OV) 0.826 75 0.049 4.424 0.583 -.139 2.308 0.849 0.569 -7.108 Table 3 Summary results of discriminant analysis evaluating the relations among fork length (FL), ovary weight (OW). plus oocyte volume lOV), for ehu (Et ?/;s carbunculus) and | kalekale iPri stipomoides sieboldii ). Anal ogous summary results using FL and OW only. Imm = immature; Mat = mature No classified No. by analysis as classified by Species histology as Imm Mat Total ': Error FL OW. plus OV Ehu Imm 33 0 33 0.00 Mat 3 145 148 2.03 Total 36 145 181 1.66 Kalekale Imm 15 0 15 0.00 Mat 1 64 65 1.54 Total 16 64 80 1.25 FL and OW Ehu Imm 33 0 33 0.00 Mat 12 136 148 8.11 Total 45 136 181 6.63 Kalekale Imm 14 1 15 6.67 Mat 3 62 65 4.62 Total 17 63 80 5.00 (ehu: 97%, kalekale: 100%) by logistic regression using FL, OW, and OV as regressor variables (Table 4). Misclassification errors increased by 3.7.5-57f (dis- criminant analysis; Table 3 ) or 0-5% ( logistic regres- sion; Table 4) if oocyte volume was excluded as a pre- dictor of maturity. Table 4 Summary statistics for logistic regressions used to clas- sify maturity (immature, mature) of ehu {Etelis carbunculus; n = l81) and kalekale iPristipomoides sieboldii; /i=80). His- tological maturity was regressed on log-transformations of three gross metrics (fork length, FL; ovary weight, OW; plus egg volume. 0\'l, and FL and OW only Intercepts and higher-order terms were tested for each model and spe- cies, but are unspecified if insignificant. Coefficient Species Regressor Estimate SE Prob Three-variable model Ehu FL ns OW -13.10 6.82 0.055 OV -2.64 0.76 0.001 OW X OV 8.03 2.58 0.002 Total df 178 Overall percentage error: 2.8'7f Kalekale FL ns OW -309.4 125.3 0.014 OV -9.55 5.26 0.069 OWxOV 125.2 47.2 0.008 Total df 77 Overall percentage error: 0% Two-variable model Ehu FL -6.08 1.46 <0.001 OW 12.29 2.75 <0.001 Total df 179 Overall percentage error; 2.8% Kalekale FL -1.74 0.61 0.004 OW 5.64 1.37 <0.001 Total df 78 Overall percentage error: 5.0% 454 Fisher/ Bulletin 97(3), 1999 20 30 40 Fork length (cm) Figure 1 Length-frequency (FL, cmi histograms of ehu. Etelis carbunculus (hollow bars, /! = 181), and kalekale, Pristipomoides sie- boldii (solid bars, n=80). Size at sexual maturity For each species, the length-frequency distribution comprised relatively few individuals <30 cm FL (Fig. 1). Hence, our size-at-maturity characterizations might have been affected by the scarcity of imma- ture specimens. Ehu On average, ehu in Hawaiian waters mature at an estimated 27.9 cm FL (95% CI=25. 5-30.3 cm) according to histological criteria or 27.8 (25.5-30.1) cm FL based on logistic regression according to gross criteria (r2=0.967 and 0.962, respectively; Fig. 2). The precision of these L^q estimates is reasonable despite the discontinuous lower tail of the size-frequency distribution. Our L^q estimates for ehu provide an adequate objective basis for evaluating gross and histological classifications; however, they should be considered only as preliminary estimates of size at maturity. Kalekale A body length-maturity pattern also was strongly evident for this species. Whether based on histological evidence or logistic regression with gross metrics, size at maturity was well described by the logistic model (r^=0.920; Fig. 3). Kalekale in Hawai- ian waters mature at 29.0 (9'5% 01=27.2-30.8) cm 100 -, c S Q- L50 = 27 8 cm FL (gross cnteria) = 27 9 cm FL (histology) Histological chtena • Px= 100/(1 + exp-(-8.247f-0,2% FL) l^ = 0 967 Gross cntena D Px- 100/(1 +ei838 mm FL from which a catch curve (Ricker, 1975) was constructed for 1987-92. We estimated instantaneous total mortality (Z) by catch curve analysis (Beverton and Holt, 1957; Everhart and Youngs. 1981) based on fully recruited fish. Results We examined 1005 cobia that ranged from 335 to 1651 mm FL, 33 of which were YOY (age 0) and ranged from 335 to 510 mm FL. External sexual di- morphism was not evident in R. canadum. Males («=275) ranged from 345 to 1450 mm FL (mean=952 mm) and from 0.3-29.0 kg (mean=10.5 kg); females (« =7301 ranged from 335 to 1651 mm FL (mean=1050 mml and from 0.3 to 62.2 kg (mean=16.6 kg). The length-frequency distributions of males and females (Fig. 3) were significantly different (Kolmogorov- Smirnov two-sample test, df =0.432, P<0.05). Females Franks et al.: Age and growth of Rachycentron canadum 463 were significantly larger than males (Mann-VVhitney f7-test, P<0.001 ), and 8591 of fish >1000 mm were female. The sex ratio of 1:2.7 was significantly different from 1:1 (x^=205.Q, df=l, P<0.6o01). Neither slopes (ANCOVA, df=914; F=2 . 156, P=0. 142 ) nor elevations ( ANCOVA, df=914. F=2.334, P=0.127) of the length- weight regressions by sex were found to be significantly different; therefore, data were pooled and one relationship estab- lished (Table 1; Fig. 4). Weight was ap- proximately a cubic function of length, implying nearly isometric growth. The re- lationships between FL and TL are pre- sented in Table 1. When viewed with transmitted light, thin-sectioned sagittae revealed a pattern of distinct, alternating narrow opaque and wide translucent bands (Fig 2). The dis- tance between the first two opaque bands distally from the core typically was wider than the distance between subsequent opaque bands. Mean marginal increment analysis (Fig. 5) demonstrated that April through August was the time of annulus formation and suggested that opaque bands form once each year. All otoliths exhibited a zone of translucent material beyond the last annulus from September through February. Mean increment was minimal during June and increased to a maximum in February (no samples were collected during December). The sample size was too small to plot marginal increments for each year and age-group separately; however, a vi- sual examination of the data indicated that marginal increments for individual years 1987-90 and age- classes 2-5 were similar, with a consistent seasonal minimum during summer. Timing of annulus forma- tion was similar for each sex. Of the 645 left sagittae processed for age estimates, 187 (29%) were judged illegible. Right sagittae from 168 of the latter group were available and processed, and 76'^?^ (128/168) were readable. Readers agreed on ages for 96% (565/586) of usable otoliths, 170 males (range 345-1330 mm FL) and 395 females (range 335-1651 mm FL). Only 21 (4%) of the us- able otoliths were rejected because of disagreements among readings, owing primarily to disparities over the presence of an annulus adjacent to the core or at the otolith's margin. Of the sagittae found accept- able for age estimations, 33 were from YOY (335- 510 mm) and 42 were from age 1 fish (493-910 mm). Ten age 1 fish were 838 mm (minimum legal size) or 80 : / 60 - J n = 915 / =' 3 5> 40 % - ,p' I ' -^' '■ °^^' ° ' ^S^'^o °° 20 _ tf^PwB^ * ° nj^° ^j^° ^^^^^[' 0 __— -°''*^ 1 1 l™f"N 1 1 1 1 1 1 1 1 1 1 1 1 1 1 1 1 1 1 1 1 1 1 1 1 1 1 1 200 400 600 800 1000 1200 1400 1600 1800 Fork length (mm) Figure 4 Length-weight relationship for cobia. Rachycentron canadum. from the northeastern Gulf of Mexico larger. Most (n=463, 82%) of the 565 fish that we aged were estimated to be ages 2-5 (27% age 2; 29%- age 3; 17% age 4; and 9% age 5). Age 6 fish and older were conspicuously uncommon. There was a signifi- cant difference between the age-frequency distribu- tions of males and females (Kolmogorov-Smirnov two-sample test, dn=0.308, P<0. 05). An age 11 female ( 1568 mm) and age 9 males (n=2, 1240 and 1260 mm) were the oldest cobia sampled (Table 2). Twenty five females (1170-1651 mm) were age 6 or older, but only six males (1035-1330 mm) were older than age 5 (Table 2). Growth in length for both sexes was relatively fast through age 2, after which growth slowed gradually (Fig. 6). We found a wide range of lengths within most age groups for both sexes (Tables 3 and 4). For ex- ample, age 4 males and females ranged from 850 to 1250 mm and from 900 to 1250 mm, respectively. We also found a wide range of ages within some of the length groups. For example, the 1000 mm and 1200 mm groups of males ranged fi-om ages 2 to 7 and ft-om 464 Fishery Bulletin 97(3), 1999 Table 1 Length-length, length-weight, and otolith weight-age re- gressions forcobia,/?ac/iycen^ror! canadiim, from the north- eastern Gulf of Mexico. FL = fork length (mm), TL = total length ( mm ), WT = total weight ( kg), OTWT = otolith weight (g), and AGE = age in years. Sample fork length range for length-length regressions and length-weight regressions was 345-1651 mm. Age range for the otolith weight-age regression was 1-9 for males and 1-11 for females. Values in parentheses are standard errors. Y = a+bX FL TL OTWT (males) OTWT (females TL FL 930 930 log „FL 915 AGE AGE 126 259 9.9494 0.8916 (3.5691) (0.0032) 1.6661 1.1088 (3.9964) (0.0040) -9.2445 3.4287 (0.6474) (0.0215) 0.0081 0.0072 (0.0012) (0.0003) 0.0006 0.0110 (0.0010) (0.0003) 0.989 0.989 0.965 0.775 0.836 Table 2 Average observed and predicted fork lengths ( mm for male and female cobia, Rachycentron canadum. Numbers in | parentheses are standard error and sample size. Males Females Age Average Average (yr) observed Predicted observed Predicted 0 439 (29.6;5) 409 (6,0;28) 1 705 (26.0;14) 709 720 (21.6;28) 713 2 885 (8.5;47) 871 956 (7.9;103) 914 3 971 (9.9:47) 976 1056 (7.2;116> 1066 4 1034 (14.2;35) 1044 1140 (10.4;64l 1183 5 1070 (16.6;16) 1089 1248 (17.6;31l 1271 6 1140 (1) 1118 1346 137. 9;7) 1339 7 1198 (86.5;3) 1136 1385 (44.0;5) 1391 8 1148 1553 (27.4;8) 1430 9 1250 (10.0;2) 1156 1507 (69.9;3) 1460 10 1613 (1) 1482 11 1568 (1) 1500 Table 3 Age-length key. Fork length (mm) composition. in percent, of male cobia by age group Length group (50 mm) Age in years Number of fish 0 1 2 3 4 5 6 7 8 9 300 100.0 1 350 400 100.0 1 450 100.0 2 500 50.0 50.0 2 550 100.0 1 600 100.0 2 650 100.0 2 700 100.0 - 4 750 100.0 1 800 11.1 83.3 5.6 18 850 4.8 66.7 19.0 9.5 21 900 40.9 50.0 9.1 22 950 20.6 50.0 29.4 34 1000 9.1 27.3 22.7 36.4 4.5 22 1050 35.3 41.2 23.5 17 1100 66.7 22.2 11 1 9 1150 50.0 50.0 4 1200 20.0 40.0 20.0 20.0 5 1250 100.0 1 1300 100.0 1 Total • 170 Franks et a\: Age and growth of Rachycentron canadum 465 ages 4 to 9, respectively (Table 3), whereas the 1350 mm group of females ranged fi'om ages 5 to 9 (Table 4). The results of likelihood-ratio tests showed a significant difference in the over- all von Bertalanffy growth models for males and females ()(-=175.06, df=l, P<0.0001) (Table 5), a finding substanti- ated by approximate randomization test- ing of the growth models {P<0.0001 ). Like- lihood-ratio tests showed that estimates of LJx^=24.60, df=l, P<0.0001) and K (X-=7.02, df=l, P=0.008) were signifantly different between sexes, however, t^ was not significantly different ( X"=-0 . 1 1 , df= 1 , P=0.752). Growth param.eters indicated that females achieved a greater theoreti- cal asymptotic length and grew at a faster rate than males. Predicted lengths-at-age derived by the von Bertalanffy equations agreed with observed lengths, except for age 9 males (n=2) and age 8 and 10 females (n-VZ) (Table 2), where observed lengths were considerably larger than those pre- dicted. Average observed lengths-at-age for females were greater than those of males for age 1 and older (Table 2), and predicted lengths of females were greater than those of males for all ages. Otolith weight was significantly related to age (Fig. 7), and the slopes of the otolith weight-age regressions for males and fe- males (Table 1) were significantly different (ANCOVA , df=385, F=34.13, P<0.0001). Age-length keys were constructed to estimate the age structure of legal-sized cobia (>838 mm FL) caught from 1987 to 92 (Fig. 8) which we believe was representative of the northeastern Gulf recreational fishery. Most (84^^) of those fish were age 2-4, whereas age 3 represented 37*^ of the catch. Age at full recruitment to the fishery was age 4 (modal age plus one). Ages 1-3 represented 66% of the fishery, age 4 represented 19%, and ages 5-11 only 15% . The instantaneous rate of total mortality (Z) estimated by our catch curve analysis for ages 4-8 was 0.75 (Fig. 9). Discussion Despite acquiring many of our cobia samples at fish- ing tournaments, we believe our overall collections reflect the recreational hook-and-line fishery for co- bia in the northeastern Gulf during the late 1980s and early 1990s. Although anglers typically enter 200 5 150 - Percent marginal increment o o - 2 1 51 27 6 50 90 34 + 1 86 1 h + 1" 56 + + + + 1 Ill 1 0 - JFIV1AMJJAS 0 N Month Figure 5 Monthly mean percent marginal increment for cobia. Rachvcentron canadum. Vertical lines represent ±1 SE. Numbers abov e vertical lines represent sample size. large fish in tournaments, substantial numbers of small fish were also entered during the competitions, particularly if aggregate weight awards were pre- sented during multiday competitions. We frequently sampled anglers' entire catch which included small fish not entered in competition. Nontournament fish were also examined at docks and marinas, and these specimens ranged from less than minimum legal size to some of the largest fish that we encountered. Although the length-weight relationships between the sexes did not differ significantly, females were typically larger than males. Thompson et al.'^ re- ported similar results for cobia taken off western Louisiana. In our study, females predominated (2.7:1 overall sex ratio) during all study years. Females were dominant in all age groups, and the magnitude of that dominance varied with increasing age. Dur- ing a five-year study (1987-91) of cobia from west- ern Louisiana waters (west of the Mississippi River delta), Thompson et al.'^ reported an overall sex ra- tio of 2.1:1 that was skewed towards males (464, 466 Fishery Bulletin 97(3), 1999 1800 1500 1200 ^J.-^ ' 900 - =/Kf"^ k length (mm) o o o o o o o _ /■ Males ■/ '■ n = 170 - o 1500 : i ■^^^^^-'r—' 1200 - -^ 1 900 !/H 600 A Females ■ y . n=395 300 0123456789 10 11 Age (years) Figure 6 Observed and predicted lengths from the von BertalanfTy growth model for male and female cobia, Rachycentron canadum. males; 218 females) for each year. Because our study and that by Thompson et al.^ were conducted con- currently, we are unable to explain this discrepancy, except to suggest differential segregation or a higher mortality for males east of the delta. Sagittal otoliths were determined to be valid age- ing structures for 7?. canadum, and alternating opaque and translucent bands were most conspicu- ous in the ventral lobe of otolith thin-sections. An- nuli were not uniformly visible in thin-sections for some fish and were occasionally obscured along the ventral sulcal ridge, particularly for fish age 5 and older. Marginal-increment analysis indicated that annuli formed once per year during April-August. Therefore, age in years for cobia was presumed equal to the number of opaque bands observed in sectioned sagittae, findings that agree with those of Thomp- son et al.'^ off Louisiana and Smith (1995) off North Carolina. Because cobia are infrequently caught in northeastern Gulf waters during the winter, the scar- city of otolith samples from November through March precluded us from making an unequivocal assertion on the annual nature of opaque band formation. However, thin-sectioned sagittae from seven cobia 0 16 - 0,14 - 0,12 Males 0 10 />= 126 0,08 - ^^ 0 06 - ^^ 0 04 ' « ^^^ f 0 02 £ ■| 0 1 0 16 o O 0 14 1 1 1 1 I 1 1 fill 0 12 Females n = 259 . y 0 10 - i/ 0 08 . . y 0 06 ■ \ yC ■ 0 04 1 V ! 0 02 / 1 1 1 1 I 1 I 1 1 1 1 0 0 12 3 4 5 6 7 8 9 10 11 Age (years) Figure 7 Sagittal weight-age relationship for ma le and female cobia, Rachycentron canadum - caught in the Florida Keys during January 1991 and sampled dockside by us showed a substantial zone of translucent material extending from the distal edge of the last opaque band to the otolith margin. This finding suggests that winter annulus formation does not occur in the otoliths of cobia from south Florida waters (cobia that may migrate into north- ern Gulf waters in spring). Although the timing of annulus formation coincides with the cobia's spawning season in the northern Gulf (Biesiot et al., 1994; Lotz et al.. 1996), annulus depo- sition may be more related to cobia migration into the northern Gulf in spring. We found that sagittae of several sexually mature cobia sampled in April (early part of the spawning season) already showed opaque bands, as did sexually immature fish in spring. The relationship of annulus formation to Franks et al,: Age and growth of Rachycentron canadum 467 Table 4 Age-length key Fork length (mm) composition, in percent, of female cobia by age group Length Age in years group Number i50 mm) 0 1 2 3 4 5 6 7 8 9 10 11 of fish 300 100.0 1 350 100.0 8 400 100.0 17 450 66.7 33.3 3 500 550 100.0 1 600 100.0 7 650 100.0 5 700 100.0 3 750 100.0 3 800 25.0 75.0 12 850 19.0 81.0 21 900 3.3 66.7 26.7 3.3 30 950 52.2 47.8 46 1000 40.0 47.3 12.7 55 1050 12.2 53.1 28.6 6.1 49 HOG 13.2 39.5 44.7 2.6 38 1150 46.4 28.6 21.4 3.6 28 1200 28.6 42.8 28.6 21 1250 40.0 40.0 10.0 10.0 10 1300 33.4 50.0 8.3 8.3 12 1350 60.0 20.0 20.0 5 1400 28.6 57.1 14.3 7 1450 50.0 50.0 2 1500 25.0 50.0 25.0 4 1550 100.0 1 1600 60.0 20.0 20.0 5 1650 100.0 1 Total 395 Table 5 Parameter estimates for the von Bertalanffy growth model for cobia, Rachycentron canadum from U.S. waters. Values shown in parentheses are standard errors. — = not reported by author(s). Area Sex n i„ K to r^ Structure Authors Virginia' M 121 0.28 -0.06 scales Richards, 1967 F — 164 0.23 -0.08 North Carolina' M 116 105 (1.85) 0.37 (0.04) -1.08 (0.29) — otoliths Smith, 1995 F 92 135 0.24 -1.53 Western Louisiana- M (3.82) 1,132 (0.03) 0.49 (0.39) -0.49 otoliths Thompson et al.^ F — 1,294 0.56 0.11 Northeastern Gulf M 170 1.170.7 0.432 -1.150 0.78 otoliths This study of Mexico- (28.081 (0.046) (0.173) F 395 1,555.0 (35.14) 0.272 (0.017) -1.254 (0.092) 0.87 ' L_ estimates reported in centimeters. - L^estimates reported in millimeters. ■' See Footnote .3 in te.'it for this source. 468 Fishery Bulletin 97(3), 1999 migration has been suggested for swordfish (Berke- ley and Houde, 1983;Tserpes and Tsimenides, 1995) and Atlantic bluefin tuna (Compean-Jimenez and Bard, 1983). Other authors (Nelson and Manooch, 1982; Sturm et al., 1989; Beckman et al., 1990; Ferreira and Russ, 1994) also suggested that repro- duction may not be the sole determining factor and commented on the physiological nature of annulus for- mation and the importance of environmental factors. Longevity of male and female cobia differed con- siderably. Males older than age 7 were rare, and maximum age was 9. Females older than age 8 were rare, and maximum age was 11. Maximum ages of cobia from Louisiana (age 10, Thompson et al.'^) and Virginia (age 10, Richards, 1967) were similar to our observations. However, Smith (1995) reported a maximum age of 14 for males and age 13 for females for cobia from North Carolina. We also found, as did Richards (1967) and Smith (1995), that mean ob- served lengths at age for females were larger than those for males for all age classes, except age 0 fish. 400 300 CD n E 200 100 n Males Females mm. \r/y7i 10 Age (years) Figure 8 Age structure of cobia, Rachycentron canadum, >838 mm FL (regula- tion minimum size) in the northeastern Gulf of Mexico recreational hook-and-line fishery, 1987-92. n=992 Considerable variation in size was observed within most age gi-oups, including YOY, for both males and females, which, according to Goodwin and Johnson (1986), is not unusual for warm-water fishes. The variation in size makes it difficult to estimate pre- cisely the age of cobia from length alone. For example, our largest cobia weighed 62.2 kg, which was slightly greater than the all-tackle world record weight for cobia (61.5 kg) reported by the International Game Fish Association ( 1997 ). At a fork length of 1610 mm and at age 8. this specimen was neither the longest fish in our sample nor the oldest. A prolonged spawn- ing season and multiple spawnings characteristic of cobia (Lotz et al., 1996) probably account for the wide variation in size of YOY cobia and other age groups as well. Annual growth was most rapid through age 2 for both sexes, then gradually decreased thereaf- ter, particularly for females. Otolith weight was a good predictor of age, ac- counted for 78*^ and 84*^ of the variability in age of male and female cobia, respectively, and explained as much variation in age as fork length in the von Bertalanffy model for each sex. Our estimates of growth parameters are the only estimates available for R. cana- dum in the northeastern Gulf We found that the von Bertalanffy theoretical growth models for males and females were signifi- cantly different, as did Thompson et al.'^ Lengths predicted from the theoretical growth curves agreed with the average observed lengths. Theoretical asymptotic lengths seemed realistic, even though few fish >1200 mm were sampled. Theoretical growth coefficients (L and t^) reported by Thompson et al.*^ for cobia from Louisiana were smaller than our estimates ( Table 5 ), although their estimates ofK were larger, particularly for females. Asymptotic lengths for males and females taken off Virginia (Richards, 1977) were consider- ably larger than L^ values reported by Smith (1995) for cobia from North Caro- lina, values reported by Thompson et al.^ for cobia from Louisiana and our study (Table 5), although our asymptotic length for males was similar to that in Richards' ( 1967) study. The differences in estimates of growth coefficients for cobia throughout their range in U.S. waters may be due to methodological differences, e.g. sectioned otoliths ( this study) versus scales (Richards, 1967), or differences in geographical cov- erage. Nevertheless, we believe our growth parameter estimates are appropriate for Franks et aL. Age and growth of Rachycentron canadum 469 4 r 0) CT ^ 2 A y= 5.826 -0.750X r= = 0.917 -*- + — ^ - '^ - ■^ 0123456789 10 11 Age (years) Figure 9 Length-converted catch curve for cobia taken in the northeastern Gulf of Mexico recreational fishery. The solid line described by the equation ( Y=a+6X) indicates the age range used in regression estimates of instanta- neous total mortality (Z). Z is equal to the absolute value of the slope (6) of the regression line. use in assessment studies of cobia from the north- eastern Gulf. Cobia were fully recruited to the recreational fish- ery in the northeastern Gulf at age 4. Catch curve analysis predicted a Z of 0.75. A fairly broad age struc- ture and a low value for Z suggest that the north- eastern Gulf population of cobia is reasonably healthy. We believe our estimate of Z is reliable, al- though several authors (Rounsefell and Everhart, 1953; Johnson, et al., 1983; and Manooch et al., 1987 ) caution against using catch curves to predict mor- tality for migratory pelagic species because, in part, such predictions are subject to a variety of assump- tions, including a constant recruitment and mortal- ity for each year and year class comprising a pooled data set. The popularity of cobia warrants contin- ued monitoring of population age structure and growth parameters of this valuable gamefish in the northern Gulf. Acknowledgments We are indebted to the anglers who allowed us to sample their catch of cobia. We thank Barbara Palko, formerly of the National Marine Fisheries Service (NMFS), Panama Citv (Florida) Laboratory, for mak- ing numerous specimens available to us. We express our appreciation to Thomas Mcllwain of the NMFS and Richard Leard of the Gulf of Mexico Fishery Management Council for their encouragement and advocacy of our work, particularly in the initial stages of the study. Many thanks to our colleagues in the Marine Fisheries Division of the Mississippi Depart- ment of Marine Resources. Chuck Wilson, Bruce Thompson, and Louise Stanley of Louisiana State University, Coastal Fisheries Institute, provided valuable advice and great inspiration. We express our sincere gratitude to Mike Allen and Dyan Wil- son for their diligence at the otolith saw and their assistance in reading otoliths. We acknowledge the many contributions of T. J. Becker, the archetypal tournament sampler. We thank Robin Overstreet for photographing the otoliths shown in this manuscript and for sharing with us his interest in the biology of cobia over many years. Many other individuals par- ticipated in portions of this study including the fol- lowing Gulf Coast Research Laboratory personnel: Don Barnes, Lisa Engel, Dale Fremin, Nikola Garber, Nate Jordan, David Lee, Jeffery Lotz, Terry McBee, Casey Nicholson, Steve Vanderkooy, and Mike Zuber. We also offer thanks to Michael Murphy and Roy Crabtree of the Florida Marine Research Institute and to James Duffey of the Alabama Department of Conservation and Natural Resources for their val- ued advice and assistance on statistical treatment of the data. We recognize colleagues Patricia Biesiot of the University of Southern Mississippi, and Joe Smith and John Merriner of the NMFS Beaufort (North Carolina) Laboratory who share with us a deep appreciation for this great fish. We thank three anonymous reviewers and the scientific editor for their extremely helpful comments and suggestions. This work was supported in part by funding fi'om Federal Aid in Sport Fish Restoration, Department of the Inte- rior, U S. Fish and Wildlife Service, Atlanta, GA, Project No. F-91, and the Mississippi Department of Marine Resources, Biloxi, Mississippi. Literature cited Beamish, R. J., and G. A. MacFarlane 1983. The forgotten requirement for age validation in fish- eries biolog\'. Trans. Am. Fish. Soc. 112:73.5-743. Beckman, D. W., A. L. Stanley, J. H. Render, and C. A. Wilson. 1990. Age and growth of black drum in Louisiana waters of the Gulf of Mexico. Trans. Am. Fish. Soc. 119:.537-544. Berkeley, S. A., and E. D. Houde. 1983. Age determination of the broadbillswordfish.X(p/i;as gladius. from the Straits of Florida, using anal fin spine sections. U. S. Dep. Commer, NOAA Tech. Rep. NMFS 8:137-143. 4 70 Fishery Bulletin 97(3), 1999 Beverton, F. J. H., and S. J. Holt. 1957. On the dynamics of exploited fish populations. Fish. Invest. Minist. A^ic, Fish. Food (G. B.). Ser. II, 19. 533 p. Biesiot, P. M., R. M. Caylor, and J. S. Franks. 1994. Biochemical and histological changes during ovarian development of cobia, Rachycentron canadum. from the northern Gulf of Mexico. Fish. Bull. 92:686-696. Briggs, J. C. 1958. A list of Florida fishes and their distribution. Bull. Fla. State Mus.. Biol. Sci. 2:221- 318. 1960. Fishes of worldwide (circumtropical) distribution. Copeia 1960(3):171-180. Cerrato, R. M. 1990. Interpretable statistical tests for growth comparisons using parameters in the von Bertalanffy equation. Can. .J. Fish. Aquat. Sci. 47:1,416-1,426. Compean-Jimenez, G., and F. X. Bard. 1983. Growth increments on dorsal spines of eastern At- lantic bluefin tuna, Thunnus thynnus, and their possible relation to migration patterns. In E. D. Prince and L. M. Pulos (eds. ), Proceedings of the international workshop on age determination of oceanic pelagic fishes: tunas, billfishes and sharks, p.111-115. U.S. Dep. Commer, NOAA Tech. Rep. NMFS 8. Crabtree, R. E., C. W. Harnden, D. Snodgrass, and C. Stevens. 1996. Age, growth and mortality of bonefish, A/6u/a vulpes. from the waters of the Florida Keys. Fish. Bull. 94:442- 451. Dawson, C. E. 1971. Occurrence and description of prejuvenile and early juvenile Gulf of Mexico cobia, Rachycentron canadum. Copeia 1960(3):171-180. Ditty, J. G., and R. F. Shaw. 1992. Larval development, distribution, and ecology of co- bia, Rachycentron canadum (Family:Rachycentridae), in the northern Gulf of Mexico. Fish. Bull. 90:668-677. Everhart, H. W., and W. D. Youngs. 1981. Principles of fishery science. Cornell Univ. Press, Ithaca, New York, NY, 349 p. Ferreira, B. P., and G. R. Russ. 1994. Age validation and estimation of growth rate of the coral trout, Ptectropomus leopardus (Lacepede 18021. from Lizard Island, Northern Great Barrier Reef Fish. Bull. 92:46-57. Franks, J. S. 1995. A pugheaded cobia (Rachycentron canadum ) from the northcentral Gulf of Mexico. Gulf Res. Rep. 9<2):143-145. Franks, J. S., N. M. Garber, and J. R. Warren 1996. Stomach contents of juvenile cobia, Rachycentron canadum, from the northern Gulf of Mexico. Fish. Bull. 94:374-380. Franks, J. S., M. H. Zuber, and T. D. Mcllwain. 1991. Trends in seasonal movements of cobia, flac/iycen^ron canadum, tagged and released in the northern Gulf of Mexico. J. Miss. Acad. Sci. 36(1 ):55. Goodwin, J. M., and Johnson A. G. 1986. Age, growth, and mortality of blue runner, Caranx crysos, from the northern Gulf of Mexico. Northeast Gulf .Sci. 8(21:107-114. Hassler, W. W., and R. P. Rainville. 1975. Techniques for hatching and rearing cobia, Rachy- centron canadum, through larval and juvenile stages. Univ. N.C. Sea Grant Coll. Prog., UNC-SG-75-30, Raleigh, NC, 26 p. Helser, T. E. 1996. Growth of silver hake within the U..S. continental shelf ecosystem of the northwest Atlantic Ocean. J. Fish. Biol. 48:1,059-1,073. Howse, H. D., J. S. Franks, and R. F. Welford. 1975. Pericardial adhesions in the cobia (Rachycentron canadum) (Linnaeus). Gulf Res. Rep. 5(11:61-62. Howse, H. D., R. M. Overstreet, W. E. Hawkins, and J. S. Franks. 1992. Ubiquitous perivenous smooth muscle cords in vis- cera of the teleost Rachycentron canadum, with special emphasis on liver J. Morphol. 212:175-189. Hrincevich, A. W. 1993. Analysis of cobia Rachycentron canadum population structure in the northern Gulf of Mexico using mitochon- drial DNA. M.S. thesis, Univ. Southern Miss., Hatties- burg, MS, 91 p. International Game Fish Association. 1997. World record game fishes. International Game Fish Association, Pompano Beach, Florida, 352 p. Johnson, A. G., W. A. Fable, M. L. Williams, and L. E. Barger. 1983. Age, growth, and mortality of king mackerel, Scomberomorus cavalla, from the southeastern United States. Fish. Bull. 81( 1 1:97-106. Joseph, E. B., J. J. Norcross, and W. H. Massmann. 1964. Spawning of the cobia, Rachycentron canadum, in the Chesapeake Bay area, with observations of juvenile specimens. Chesapeake Sci. 5:67-71. Kimura, D. K. 1980. Likelihood methods for the von Bertalanffy growth curve. Fish. Bull. 77(4):765-776. Knapp, F. T. 1949. Menhaden utilization in relation to the conservation I of food and game fishes of the Texas Gulf Coast. Trans. I Am. Fish. Soc. 79:137-144. 1 Knapp, F. T. ( 1951. Food habits of the sergeantfish, Rachycentron canadus. Copeia 1951:101-102. Lotz, J. M., R. M. Overstreet, and J. S. Franks. 1996. Gonadal maturation in the cobia, Rachycentron canadum. from the northcentral Gulf of Mexico. Gulf Res. Rep. 9(3):147-159. Manooch, C. S., S. P. Naughton, C. B. Grimes, and L. Trent. 1987. Age and growth of king mackerel, Scomberomorus cavalla, from the U. S. Gulf of Mexico. Mar Fish. Rev. 49(21:102-108. Meyer, G. H., and J. S. Franks. 1996. Food of cobia, Rachycentron canadum , from the northcentral Gulf of Mexico. Gulf Res. Rep. 9(3):161-167. Miles, D. W. 1949. A study of the food habits of the fishes of the Aransas Bay area. M.S. thesis, Univ. Houston, TX, 70 p. Nelson, R. S., and C. S. Manooch. 1982. Growth and mortality of red snappers in the west- central Atlantic Ocean and northern Gulf of Mexico. Trans. Am. Fish. Soc. 111:465-475. Richards, C. E. 1967. Age, growth and fecundity of the cobia, Rachycentron canadum. from the Chesapeake Bay and adjacent Mid- Atlantic waters. Trans. Am. Fish. Soc. 96:343-350. 1977. Cobia (Rachycentron canarfz/m) tagging within Chesa- peake Bay and updating of growth equations. Chesapeake Sci. 18:310-311. Rickcr, W. E. 1975. Computations and interpretation of biological statis- tics offish populations. Fish. Res. Board Can., Bull. 191, 382 p. Franks et al : Age and growth of Rachycentron canadum 471 Rounsefell, G. A., and W. H. Everhart. 1953. Fishery science: its methods and applications. John Wiley and Sons, Inc.. New York, NY, 444 p. Secor, D. H., J. M. Dean, and E. H. Laban. 1992. Otolith removal and preparation for microstructural examination. In D. K. Stevenson and S. E. Campana (eds.). Otolith microstructure examination and analysis, p. 19-57. Can. Spec. Publ. Fish. Aquat. Sci. 117. Shaffer, R. V., and E. L. Nakamura. 1989. Synopsis of biological data on the cohia Rachycentron canadum (Pisces: Rachycentridaet. FAO Fisheries Synop. 153 (NMFS/S 1531. U.S. Dep. Commer., NOAATech. Rep. NMFS 82, 21 p. Smith, J. W. 1995. Life history of cobia. Rachycentron canadum (Osteich- thyes:Rachycentridae), in North Carolina waters. Brim- leyana 23:1-23. Snedecor, G. W. and W. C. Cochran. 1967. Statistical methods. 6th ed. Iowa State Univ. Press, Ames, lA, 593 p. Statgraphics 1994. Statistical graphics software, ver. 7.1. Manugistics, Inc., Rockville, MD. Sturm, M. G. de L., and P. Salter. 1989. Age, growth and reproduction of the king mackerel Scomberomorus cavalla iCuvier) in Trinidad waters. Fish. Bull. 88:361-370. Tserpes, G., and N. Tsimenides 1995. Determination of age and growth of swordfish, Xiphias gtadius L., 1758, in the eastern Mediterranean using anal-fin spines. Fish. Bull. 93:594—602. von Bertalanffy, L. 1957. Quantitative laws in metabolism and growth. Quart. Rev Biol. 32:217-231. Wilson, C. A., R. J. Beamish, E. B. Brothers, K. D. Carlander, J. M. Casselman, J. M. Dean, A. Jerald, E. D. Prince, and A. Wild. 1987. Glossary. In R. C. Summerfelt and G. E. Hall (eds.), Age and growth of fish, p. 527-530. Iowa State Univ. Press, Ames, Iowa. 472 Abstract.— We examined scale samples from historical collections of postsmolts from the Gulf of St. Lawrence, Canada, with the aim of understanding the role of estuarine and coastal habitats as a juvenile nursery for Atlantic salmon. Circuli spacing patterns were extracted from the scales of 580 postsmolts col- lected in the Gulf during three seasons, 1982-84. Poststratification of the samples by collection date within year suggests that in some years postsmolts remain in the Gulf throughout the en- tire summer growth season, whereas in other years only slower growing fish re- main in these areas. Growth patterns for Gulf of St. Lawrence postsmolts were compared with patterns for re- turns from three salmon stocks from the southern end of the range in North America. These data suggest that in some years postsmolt growth in the Gulf is as robust as that observed for both the one seawinter ( ISW) and two seawinter (2SW) returns to southern rivers. Postsmolts are believed to use oceanic nursery areas generally; thus, comparable growth between the two stock groups suggests that the Gulf may serve as an important part of the postsmolt nursery range in some years. The concept of the postsmolt nursery as a continuum between neritic and oceanic areas is essential to evaluating ocean climate and productivity effects on salmonid recruitment. Growth patterns in postsmolts and the nature of the marine juvenile nursery for Atlantic salmon, Salmo salar Kevin D. Friedland UMass/NOAA CMER Program Blaisdell House, University of Massachusetts Amherst, Massachusetts 01003 E-mail address friedlandkd forwild umassedu Jean-Denis Dutil Ministere des Peches et des Oceans Institut Maurice-Lamontagne 850 route de la Mer, Mont-Joli, Quebec, Canada G5H 3Z4 Teresa Sadusky National Marine Fisheries Service, NOAA 166 Water Street Woods Hole, Massachusetts, 02543 Manuscript accepted 11 August 1998. Fish. Bull. 97:472-481 ( 1999). Anadromous salmonids, such as Atlantic salmon, Salmo salar, use freshwater stream habitats as nurs- ery areas for their early life history stages (Thorpe, 1994). Freshwater residency provides refuge from po- tential predators and relatively stable survival conditions for eggs and juveniles (Chaput et al., 1998). Thus, the recruitment of smolts during the freshwater phase is char- acterized by relatively low annual variability, most of which is attrib- uted to the size of spawning escape- ments (Chadwick, 1987). As a con- sequence, the relationships between spawning stock abundance and re- cruits migrating from freshwater take on predictable forms (Chad- wick, 1985; Elliott, 1993). The rela- tionships between generations of spawners, however, are not as clear owing to highly variable rates of marine survival. The first year at sea, or the post- smolt year, for Atlantic salmon is poorly understood. The transition to the marine environment is in itself a survival challenge associated with specific estuarine habitats (Levings, 1994). After postsmolts make the transition, they are believed to dis- perse widely in ocean surface wa- ters. This belief stems from the facts that they are rarely caught in near- shore fishing gears and have proven difficult to capture in directed sur- veys. The exception has been in the Baltic area, where the spatial con- fines of the Baltic Sea and postsmolt recoveries in commercial fisheries have combined to produce a relative wealth of information on the distri- bution and survival of postsmolts in that region (Larsson, 1985; Eriks- son 1994; Kuikka and Salminen 1994). The equivalent information does not exist for North Atlantic postsmolts; thus, it is critical to learn more about the basic biology of postsmolts so that informed in- terpretations of the patterns in salmonid population dynamics can be made. Recoveries of North American versus European postsmolts sug- gest that habitat use and ecological roles may differ for the early ma- rine stages of the two stock com- plexes. Recent captures of posts- molts in the Northeast Atlantic show that they are distributed over Fnedland et al . Growth patterns in postsmolts of Salmo solar 473 a large oceanic region by their first summer at sea (Holm et al., 1996; Shelton et al., 1997). These distribution patterns suggest that postsmolts migrate to the northwest and may be less likely to use inshore ner- itic habitats. However, in the Northwest Atlantic, the few recovery data available for postsmolts suggest differing views of which habitats represent the nursery area for postsmolts. Salmon postsmolts have been reported to use estuarine waters; therefore it is possible that they possess specific be- haviors to orient themselves to inshore ar- eas during their first year at sea ( Robitaille et al., 1986; Dutil and Coutu, 1988; Cunjak et al., 1989). In contrast, postsmolts have also been collected from research surveys in the Labrador Sea and indirectly from bird colonies, providing e\idence that the extent of the postsmolt migration includes waters north of Newfoundland ( Montevecchi et al., 1988; Reddin and Short, 1991). Recoveries of postsmolts in estuarine areas support the hypothesis that specific habitats serve as nursery areas for postsmolts, whereas recov- eries in the pelagic zone support the alterna- tive, that juveniles have thei'mal preferences similar to adults and do not use specific postsmolt nursery areas. In this paper we re-examined material collected by Dutil and Coutu (1988) with the aim of under- standing the use of the Gulf of St. Lawrence by salmon postsmolts. Specifically, we collected circuli spacing data from salmon scales to address the fol- lowing questions: 1) Are postsmolts retained in the Gulf of St. Lawrence during the postsmolt growth season or are they transient in the area?; and 2) Do postsmolts recovered in the Gulf exhibit growth pat- terns similar to postsmolts from other stocks? The answers to these questions will hopefully allow us to make some inference about the nature of the nurs- ery for Atlantic salmon and salmonids in general. Figure 1 Map of Northwest Atlantic area showing postsmolt habitat for stocks in the Gulf of St. Lawrence and south (horizontal hatchingi and sampling area for Gulf of St. Lawrence postsmolts (vertical hatching). Table 1 Sample sizes for circu i anal. v'sis by stock group and age of capture. Stock Smolt year Age 1982 1983 1984 Connecticut 2SW 52 81 110 Penobscot ISW 65 63 57 2SW 75 75 40 St. John ISW 84 65 51 2SW 81 67 54 Gulf of St. La wrence Post ^molt 372 154 54 Material and methods We collected scale circuli spacing data representa- tive of postsmolt gi'owth for juvenile salmon captured in the Gulf of St. Lawrence and for thi'ee index stocks froin the southern portion of the range of salmon in North America. Gulf of St. Lawrence postsmolt salmon were collected in 1982-84 and were originally reported in Dutil and Coutu ( 1988 ). These postsmolts were captured in experimental gill nets along the northwest shore of the Gulf during the months from August to October (Fig. 1). Data for comparative purposes come from the returning adults of hatch- ery origin fish from the Connecticut, Penobscot, and Saint John rivers, all located south of the Gulf of St. Lawrence (Fig. 1). Data for the Connecticut and Penobscot rivers have been presented earlier (Fried- land et al., 1996b); whereas, the Saint John data are newly reported here. Sample sizes sorted by smolt year of migration to sea and age of maturity are re- ported in Table 1. The low number of ISW returns to the Connecticut River did not provide sufficient samples for inclusion in our study. 474 Fishery Bulletin 97(3), 1999 360° Axis B Post-smolt Growth Zone 360° Axis Freshwate Growth Zone Fjrestiwater 'prowth Zone Figure 2 Illustrations of Atlantic salmon scales from a ISW salmon (A) and a postsmolt salmon (B). Focus of each scale is marked; freshwater growth zone extends from focus to start of marine growth; and postsmolt growth zone extends to first winter annuli. The figures are not in scale. Postsmolt growth was interpreted from circuli spacing patterns deposited during the postsmolt year Scales were cleaned and mounted between glass slides and the spacings of scale circuli were measured with a Bioscan Optimas image processing system. The first spacing was measured between the first circulus of the postsmolt growth zone and the next circulus and continued with successive circuli pairs. For maturing fish returning to their natal river af- ter at least one winter at sea, the measurements were made through the first seawinter annulus of the scale and thus captured the entire postsmolt growth zone (Fig. 2A). For fish captured as postsmolts, measure- ments were taken to the edge of the scale (Fig. 2B). Thus, only part of the data collected from maturing fish was needed for comparison with the postsmolts. Measurements were made on a single scale from each specimen at a pixel resolution of 0.004 mm along the 360° axis of the scale. Return rates by sea-age and fraction of the smolt cohort maturing after one seawinter ( ISW) were com- puted for the three stocks used for comparison. Re- turn rates are simple percentages of the number of returns-at-age to the number of smolts released. Return rates for the Connecticut and Penobscot stocks originally reported in Friedland et al. ( 1996b) are updated here. Return rates for the Saint John stock were computed similarly and are based on smolt releases of 172, 145, and 206 thousand smolts for the years 1982-84, respectively. Releases over the period 1974-92 averaged approximately 200 thou- sand smolts annually in the Saint John system. The fraction of the smolt cohort maturing after one seawinter, the ISW fraction, was computed with the formulae of Friedland et al. ( 1996b). Likewise, these data are updated for the Connecticut and Penobscot stocks and are newly reported for the Saint John stock. Circuli spacing data for Gulf of St. Lawrence postsmolts were compared in three ways. In the first comparison, data for the postsmolt scales were com- pared for the period 1982-84. The spacings of the first ten circuli pairs were compared with ANOVA. This and subsequent statistical analyses were re- stricted to the first ten circuli pairs because most postsmolt scales did not have circuli beyond the tenth pair. As a consequence, spacing patterns for circuli pairs beyond pair number 10 were not well estimated. In the second comparison, for two of the seasons, 1982 and 1983, where sufficient samples of postsmolts were available, samples were poststratified by col- lection date. The poststratification was done to achieve nearly equal numbers of samples in each stratum. In 1982, three within-year strata were cre- ated: 8 to 17 August composed of 126 samples, 18 August to 5 September composed of 12 1 samples, and 7 September to 18 October composed of 125 samples. In 1983, two strata were created: 24 to 30 Septem- ber composed of 73 samples and 1 to 11 October com- posed of 81 samples. Circuli spacing of the first ten circuli pairs for the poststratified groups were com- pared with ANOVA. For the third comparison, cir- culi spacing data from Gulf postsmolt scales were compared with the circuli spacing data for the three Friedland et al.: Growth patterns in postsmolts of Salmo solar 475 Table 2 Results of ANOVAs testing from 1982 to 1984. year effect in scale circul spacing from co Uections of postsmolt salmon from the G ulfofSt. Lawrence Spacing pair df effect MS effect df error MS error F P-level 1 2 0.00291 577 0.00015 19.312 0.00 * 2 2 0.00023 577 0.00020 1.170 0.31 3 2 0.00112 577 0.00024 4.605 0.01 * 4 2 0.00290 577 0.00021 13.491 0.00 * 5 2 0.00591 576 0.00020 28.895 0.00 * 6 2 0.00700 575 0.00017 40.579 0.00 * 7 2 0.00620 574 0.00019 33.492 0.00 * 8 2 0.00482 573 0.00015 31.914 0.00 * 9 2 0.00398 559 0.00016 24.686 0.00 * 10 2 0,00217 526 0.00016 13.521 0.00 * comparison stocks. The data for the com- parison stocks were grouped by sea-age; thus, the postsmolt scale data were com- pared with a total of five scale growth sig- nals. The statistical comparison was sim- plified to a two-group ANOVA, namely to comparing circuli spacing of the first ten circuli pairs of the postsmolts with a com- bined sample of all the comparison stocks. Results Circuli spacing patterns for the postsmolt growth zones of Gulf of St. Lawrence postsmolts were significantly different for the three year classes we examined. Cir- culi spacing for the first few pairs, associ- ated with the first few weeks in the ma- rine environment, were similar in all three years, measuring approximately 0.050 to 0.055 mm (Fig. 3). However, beyond these initial pairs the spacings diverged into three distinct patterns. Spacings for fish captured in 1982, based on the mean spacings for pairs 4-8, were the widest, generally larger than 0.065 mm. Circuli spacings in the same region of the scale were progressively smaller for samples collected in 1983 and 1984. In 1983, most of the mean spac- ings did not exceed 0.060 mm and for 1984 they did not exceed 0.055 mm. ANOVA results show that the yearly spacings means were significantly different for all but one of the first ten spacing pairs (Table 2). The means for spacing pair number 2 is not signifi- cantly different because the rank order of the means changes between spacing pairs 1 and 3. Circuli pair Figure 3 Circuli spacing versus circuli pair for Gulf of St. Lawrence postsmolts for three smolt year classes, 1982-84. Error bars mark 95'7f confidence intervals. Within-year comparisons of Gulf of St. Lawrence samples poststratified by collection date suggest that in one year the same group or population offish were sampled during the entire growth season; whereas, in another year, the postsmolt assemblage appeared to change during the sampling period. All three poststratified groups formed from the 1982 data showed a similar pattern of circuli spacing versus circuli pair through pair number 12 (Fig. 4A). After circuli pair 12, the mean spacing for the three poststratified groups began to diverge; however, the mean spacing for these circuli pairs was estimated 476 Fishery Bulletin 97(3), 1999 with low precision owing to the small sample sizes available. No significant differences existed between the spacing means for circuli pairs 1-10 (Table 3). In contrast, the data for 1983 showed that spacing pat- terns and growth were different for early versus late season collections. Fish collected in September had wider circuli spacings for circuli pairs 4-10 than fish collected in October (Fig. 4B). The September samples had mean spacings generally above 0.0625 mm; whereas, the October samples had mean spacings generally below 0.06 mm. Most of the statistical com- parisons between the poststratified groups were sig- nificant at either P=0.05 or P=0.1 (Table 3). The seasonal pattern in circuli spacing for Gulf of St. Lawrence postsmolts was similar to the patterns observed for the some of the comparison stocks by year, and dissimilar in other cases. For smolt year 1982, Gulf of St. Lawrence postsmolts and Penobscot and Saint John returns showed similar seasonal pat- terns of circuli spacing versus circuli pair (Fig. 5A). as a. CO 0 07. 0.06- 0.05- 0.04- 0.07- 0,06- 0.05- 0.04. -o— Aug 8-17 -A— Aug 18-Sept 5 -V— Sept 7-Oct 18 -.— I 1 1 1 I I I — < I ' I < I ' I < I — ' — I — r— r B Sept 24-30 Oct 1-11 T 1 1 1— I— 1 1 I < I r— I r— I 1— p-i 1— I— I r— f- 0 2 4 6 8 10 12 14 16 18 20 22 Circuli pair Figure 4 Circuli spacing versus circuli pair for Gulf of St. Lawrence postsmolts for two smolt year classes, 1982 (A) and 1983 (B I. Samples are poststratified by date of capture. Error bars mark 95'7f confidence intervals. This comparison included both the ISW and 2SW maturity groups within a stock. The pattern of mean circuli spacing for Connecticut River 2SW returns was different from that for the other stocks and showed an ontogenetic trajectory of narrower spac- ings with age. When compared statistically, mean spacings for Gulf postsmolts were significantly higher and lower than the mean for the three comparison stocks (Table 4). The mean spacing for circuli pair 1 was significantly lower for Gulf postsmolts compared with the other stocks, but the relationship between stock circuli spacing changed for pairs 4-8 as the rank order changed. The rank order changed again by circuli pair 10. In 1983, the spacing pattern for the postsmolts could not be distinguished graphically from the combined signals for the other stocks (Fig. 5B). which is supported by statistical comparison as well (Table 4). Again, the Connecticut River stock had the narrowest spacings, whereas the widest spac- ings were observed in Penobscot and Saint John fish. In 1984, mean circuli spacing varied over a wide range, suggesting the different stocks experienced different growth re- gimes that year (Fig. 5C). What is par- ticularly striking is that the wild-origin postsmolts from the Gulf had the narrow- est mean spacings for most of the post- smolt growth season. Mean spacings for Gulf postsmolts were significantly lower than those for the comparison stocks with the exception of only one spacing pair (Table 4). The three stocks provided for compari- sons with Gulf of St. Lawrence postsmolts displayed a range of return rates and ISW fraction, reflecting differences in survival and maturity rates. For the Con- necticut stock, return rates of 2SW fish and ISW fraction averaged 0.13% and 0.01, respectively (Table 5). Penobscot River smolts returned at higher rates, 0.097^ and 0.477f for ISW and 2SW fish, respectively. Typically 10*^ of the Penob- scot cohort matured at ISW. The Saint John stock had the highest return rates and the largest proportion of the cohort maturing after one-seawinter. The return rate for ISW fish averaged lAT/r and the ISW fraction averaged 0.59. The 2SW return rate for the Saint John stock is similar to that observed for the Penobscot stock. Thus, the Connecticut stock had the lowest survival rate and produced few early maturing fish; the Saint John stock had the highest survival rate and pro- Friedland et al.: Growth patterns in postsmolts of Salmo salar 477 duced many early maturing fish; and the Penobscot stock was intermediate between the other two. Discussion Our analysis suggests that the role of the Gulf of St. Lawrence as salmon postsmolt nursery habitat var- ies annually. In some years it appears that the growth of wild postsmolts retained in the Gulf is as robust as and patterned similarly to growth obsei^ved for hatchery-origin postsmolts assumed to use open ocean habitats (Reddin and Short, 1991). This corre- lation suggests that either postsmolts from other areas invade the Gulf and use it as a nursery area or the Gulf region is continuous with a larger area of similar growth conditions where the nursery is formed. In other years, it appears that only smaller, and presumbably less robust, postsmolts remain in the Gulf area and that the nursery was formed else- where. Variation in the suitability of nursery habi- tat must interact with the ability of postsmolts to migrate successfully to more favorable areas. In what was first investigated as a mechanism controlling the return migration of adults. Groot and Cooke ( 1987 ) described an analogous situation with the dis- tribution of juvenile sockeye salmon in the Strait of Georgia. It would appear that the dominant wind patterns in the Strait can alter the annual migra- tion of postsmolt sockeye salmon and place them in different nursery areas each year (Peterman et al., 1994). Atlantic salmon postsmolt migration trajec- tories may be similarly affected. In some years, postsmolts are deposited in areas like the Gulf of St. Lawrence and growth conditions are favorable enough to retain them in the region for the entire season. In other years they are not deposited or re- tained in the same area. Regardless, our focus shifts to the factors controlling migration and whether they covary with the factors controlling sui-vival. We view nursery habitat for postsmolt salmon as being dynamically defined because it shifts spatial location on an annual basis to regions where the pro- duction will support growth. Many marine fishes use staged distribution separations between estuarine, coastal, and offshore habitats (Blabber et al., 1995). North American origin salmon are generally concen- trated in the Labrador Sea as feeding adults or on various migration routes back to their natal rivers as maturing fish (Reddin and Shearer, 1987). How- ever, postsmolt distributions are regulated in part by passive displacement mechanisms and the swim- Table 3 Resu of St. ts of ANOVAs comparing scale circuli spacing Lawrence during 1982 and 1983. for successive time periods from collections of postsmolt salmon from the Gulf Year Spacing pair df effect MS effect df error MS error F P-level 1982 1 2 0.00015 369 0.00014 1.085 0.34 2 2 000008 369 0.00021 0.365 0.69 3 2 0.00036 369 0.00022 1.612 0.20 4 2 0.00013 369 0.00021 0.639 0.53 5 2 0.00021 369 0.00018 1.167 0.31 6 2 0.00002 369 0.00015 0.137 0.87 7 2 0.00038 368 0.00014 2.655 0.07 8 2 0.00003 368 0.00013 0.215 0.81 9 2 0.00017 355 0.00014 1.202 0.30 10 2 0.00027 324 0.00015 1.849 0.16 1983 1 0.00001 152 0.00013 0.057 0.81 2 0.00000 152 0.00015 0.000 0.99 3 0.00014 152 0.00024 0.582 0.45 4 0.00120 152 0.00018 6.535 0.01 * 5 0.00133 152 0.00024 5.424 0.02 6 0.00129 152 0.00018 6.977 0.01 * 7 0.00084 152 0.00027 3.067 0.08 8 0.00129 152 0.00019 6.713 0.01 9 0.00064 152 0.00021 3.076 0.08 10 0.00093 152 0.00017 5.412 0.02 * »78 Fishery Bulletin 97(3), 1999 ming potential of the fish (Reddin and Friedland, 1993). As such, postsmolt mi- gration routes are unlikely to be equiva- lent among years (Caron, 1983; Jonsson et al., 1993). At some point during the postsmolt's first growing season, swim- ming ability begins to exceed current ve- locity and postsmolts can more effectively modify their distribution according to preferences driven by migi-ation mecha- nisms or foraging behavior. These factors may allow postsmolts to concentrate in specific habitats that best suit their feed- ing requirements and afford them some measure of protection from predation. However, the process of habitat selection may result in a nursery that encompasses different areas each year and thus not linked to a specific area (Friedland et al., 1996a). Therefore, years of poor feeding and growth conditions in the Gulf of St. Lawrence do not necessarily mean poor sui'vival conditions because they do not preclude the use of other neritic areas as postsmolt nursery. For example, in some years the nursery may form along the south coast of Newfoundland or for that matter make use of few neritic habitats as the fish distribute in offshoi'e areas. Contemporary characterizations of postsmolt populations may be inadequate if they fail to account for the distribution of postsmolts in neritic habitats. Coher- ence patterns in the performance of stock complexes has led investigators to search for broad-scale forcing functions to ex- plain variation in returns rate by age (Friedland et al., 1993; Friedland et al., 1998a). In the Northeast Atlantic, a direct link between postsmolt survival and ocean climate has been reported (Friedland et al., 1998a). Spring thermal conditions were associated with the survival of North Sea stocks, which argues that the postsmolt nursery in that region was oce- anic and thus directly affected by ocean climate change. However, interpretation of survival signals in the Northwest Atlantic has been complicated by the interplay of mortality and maturation (Friedland et al., 1998b). As the search continues for the factors affecting the survival of Northwest Atlantic postsmolts, investigators must be cognizant of the fact that much of the nursery may occur in inshore neritic waters rather than in pelagic ecosystems and thus may not respond solely to ocean-scale phenom- ena. This situation may be particularly important in evaluating coastal and offshore predators and how 0.07- 0.06 K 0.05 0.04- 0.04 «*^1 i+f-H^N^^fc;) H "T "T 0 2 4 6 8 10 12 14 16 18 20 22 Circuli pair Figure 5 Circuli spacing versus circuli pair for Connecticut 2SWiConnl, Penobscot ISW and 2SW (Pen), Saint John ISW and 2SW (SJl, and Gulf of St. Lawrence postsmolts (Gulf) for three sniolt year classes, 1982 (A), 1983 (B), and 1984 iC). Error bars mark 9.5'^; confidence intervals. they might be impacting salmon populations. For example, a gannet colony may exert predation pres- sure over a substantial part of the northern coast of North America; thus, these and other bird species that are adapted to surface prey, such as Atlantic salmon postsmolts, could cause significant predation (Montevecchi and Myers, 1997). The Gulf of St. Lawrence is a large, complex sys- tem that offers a diversity of feeding conditions to smolts upon their entrance into the marine environ- ment. Growth rates would be expected to vary spa- tially in the Gulf owing to, among other things, highly variable thermal conditions over years. Nursery habi- tats within the Gulf appear to provide conditions that would support growth rates comparable to those ob- served in hatchery-reared stocks. But, what clearly differentiates the Gulf habitats from most offshore areas is the rapid continental cooling that occurs in the fall, resulting in an emigration of postsmolts from Fnedland et al.: Growth patterns in postsmolts of Salmo salar 479 Table 4 Results of ANOVAs comparing scale circuli spacing for Gulf of St. he wrence pos smolts and ISW and 2SW ret urns to the Con- necticut, Penobscot, and St. John rivers for smolt years 1982-84. Year Spacing pair df effect MS effect df error MS error F P-level 1982 1 1 0.00702 727 0.00015 47.053 0.000 * 2 1 0.00057 727 0.00018 3.155 0.08 3 1 0.00086 727 0.00020 4.364 0.04 * 4 1 0.00128 727 0.00018 7.165 0.01 * 5 1 0.00425 727 0.00017 24.541 0.00 * 6 1 0.00256 727 0.00015 16.600 0,00 * 7 1 0.00248 726 0.00015 16.517 0.00 * 8 1 0.00061 726 0.00015 4.086 0.04 * 9 1 0.00001 713 0.00015 O.OSl 0.78 10 1 0.00081 682 0.00017 4.872 0.03 * 1983 1 1 0.00050 503 0.00016 3.117 0.08 2 1 0.00000 503 0.00019 0.004 0.95 3 1 0.00001 503 0.00021 0.029 0.87 4 1 0.00001 503 0.00023 0.032 0.86 5 1 0.00005 503 0.00022 0.232 0.63 6 1 0.00003 503 0.00020 0.156 0.69 7 1 0.00018 503 0.00021 0.873 0.35 8 1 0.00002 503 0.00018 0.093 0.76 9 1 0.00006 503 0.00020 0.277 0.60 10 1 0.00000 503 0.00020 0.002 0.97 1984 1 1 0.00168 364 0.00018 9.118 0.00 * 2 1 0.00125 364 0.00020 6.302 0.01 * 3 1 0.00052 364 0.00022 2.396 0.12 4 1 0.00327 364 000020 11.568 0.00 * 5 1 0.00336 363 0.00020 16.552 0.00 * 6 1 0.00552 362 0.00022 25.136 0.00 * 7 1 0.00551 362 0.00015 35.738 0.00 * 8 1 0.00606 361 0.00016 39.032 0.00 * 9 1 0.00783 360 0.00016 48.985 0.00 * 10 1 0.00512 358 0.00017 31.024 0.00 * Table 5 Percent return rate by age group and ISW fraction for three hatchery index stocks. Return rates are in proportion of cohort maturing after one sea-winter. Mean is for the period smolt years 1974-92. percent where fraction is Smolt year Connecticut Penobscot St. John 2SW Fraction ISW 2SW Fraction ISW 2SW Fraction 1982 0.02 0.00 0.05 0.52 0.08 0.91 0.65 0.49 1983 0.30 0.02 0.05 0.66 0.06 1.00 0.61 0.54 1984 0.09 0.00 0.04 0.59 0.04 0.98 0.39 0.59 Mean 0.13 0.01 0.09 0.47 0.12 1.47 0.57 0.59 480 Fishery Bulletin 97(3), 1999 the region (Dutil and Coutu, 1988). The northern Gulf can cool to temperatures below 2°C by late Novem- ber and to lethal temperatures later in winter, from the surface to a depth of 100 m. Thus, the Gulf must be viewed as a seasonally transient habitat for postsmolts in most, if not all, years. Our comparisons of growth patterns could be en- hanced with the inclusion of growth information from stocks originating in the Gulf of St. Lawrence itself We assumed that the range of growth patterns pro- vided by the three comparison stocks is representa- tive of stocks outside the Gulf region. However, the samples failed to account for growth signals from wild stocks and included only populations from the south- ern portion of the range. However, the strength of the sample is that it did include three stocks of varying growth and survival characteristics, thus supporting the contention that it likely represents a range of growth responses similar to those occurring in wild stocks. A future challenge for research would be to char- acterize the thermal properties and production char- acteristics of the area potentially comprising the postsmolt nursery to evaluate its spatial and tempo- ral extent and annual variability. From these descrip- tive analyses, it may be possible to design migration simulations and field experiments that could begin to describe the mechanisms that form the postsmolt nursery and, in turn, it may be possible to develop an understanding of the factors affecting postsmolt survival and recruitment in salmonid populations. Acknowledgments We thank L. Marshall, Department of Fisheries and Oceans, Halifax, for providing scale material from the Mactaquac Hatchery and summary return rate data, and J. Kocik and J. Boreman for reviewing early drafts of the manuscript. Literature cited Blabber, S. J. M., D. T. Brewer, and J. P. Salini. 1995, Fish communities and the nursery role of shallow inshore water of a tropical bay in the Gulf of Carpentaria, Australia, Estuarine Coastal Shelf Sci, 40(21:177-9.3. Caron, F. 1983. Migration toward the Atlantic of postsmolt iSalmo salar) from the (lulf of St, Lawrence, Nat, Can, 110(2): 223-227. Chadwick, E. M. P. 1985, The influence of spawning stock on production and yield of Atlantic salmon, Salmo salar L,, in Canadian rivers, Aquacult. Fish. Manage, 16(11:111-719. 1987. Causes of variable recruitment in a small Atlantic salmon stock. Am. Fish. Soc. Symp, Ser. 1:390-401. Chaput, G., J. Allard, F. Caron, J. B. Dempson, C. C. Mullins, and M. F. O'Connell. 1998. River-specific target spawning requirements for At- lantic salmon iSalmo salar) based on a generalized smolt production model. Can, J, Fish. Aquat, Sci. 55:246-61. Cunjak, R. A., E. M. P. Chadwick, and M. Shears. 1989. Downstream movements and estuarine residence by Atlantic salmon parr uSahito salar). Can, J, Fish. Aquat. Sci, 46(91:1466-1471, Dutil, J.-D., and J.-M. Coutu. 1988, arly marine life of Atlantic salmon, Salmo salar, post- smolts in the northern Gulf of St. Lawrence. Fish, Bull. 86(21:197-212. Elliott, J. M. 1993. A 25-year study of production of juvenile sea-trout, Salmo triitta. in an English Lake District stream. Can. Spec, Publ. Fish. Aquat, Sci, 118:109-22. Eriksson, T. 1994. Mortality risks of Baltic salmon during downstream migration and early sea-phase: effects of body size and season. Nord. J. Freshwater Res. 69:100, Friedland, K. D., D. W. Ahrenholz, and J. F. Guthrie. 1996a, Formation and seasonal evolution of Atlantic men- haden juvenile nurseries in coastal estuaries. Estuaries 19(11:105-114. Friedland, K. D., R. E. Haas, and T. F. Sheehan. 1996b, Postsmolt growth, maturation, and survival of two stocks of Atlantic salmon. Fish, Bull. 94(41:654-63, Friedland, K. D., L. P. Hansen, and D. A. Dunkley. 1998a. Marine temperatures experienced by postsmolts and the survival of Atlantic salmon (Salmo salar L.) in the North Sea area. Fish. Oceanogr, 7:22-34. Friedland, K. D., D. G. Reddin, and J. F. Kocik. 1993. Marine sun'ival of North American and European Atlantic salmon: effects of growth and environment. ICES J. Mar Sci. 50:481-92. Friedland, K. D., D. G. Reddin, N. Shimizu. R. E. Haas, and A. F. Youngson. 1998b, Strontium:calcium ratios in Atlantic salmon otoliths and observations on gi'owth and maturation. Can. J. Fish. Aquat, Sci. 55:1158-1168, Groot, C, and K. Cooke. 1987, Are the migrations of juvenile and adult Fraser River sockeye salmon iOncorhynchus nerka) in near-shore wa- ters related? Can. Spec. Publ, Fish. Aquat. Sci. 96: 53-60. Holm, M., J. Hoist, and L. Hansen. 1996. Atlantic salmon surveys in the Norwegian Sea from •July 1991-August 1995. Aquaculture 1996(1 ):21. Jonsson, N., L. P. Hansen, and B. Jonsson. 1993. Migratory behavior and growth of hatchery-reared postsmolt Atlantic salmon Salmo salar J. Fish Biol. 42:435-443. Kuikka, S., and M. Salminen. 1994. Dependency of stocking result on the size of the re- leased salmon sniolts tSalmo salar L.) in the northern part of the Baltic Sea. Nord. J. Fre.shwater Res. 69:99. Larsson, P.-O. 1985. Predation on migrating smolt as a regulating factor in Baltic salmon, Salmo salar L., populations. J. Fish Biol. 26(41:391-397. Levings, C. D. 1994. Feeding behavior of juvenile salmon and significance of habitat during estuary and early sea phase. Nord. J. Freshwater Res. 69:7-16. Fnedland et al Growth patterns in postsmolts of Salmo salai 481 Montevecchi, W. A., D. K. Cairns, and V. L. Birt. 1988. Migration of postsmolt Atlantic salmon. Salmo salar. off northeastern Newfoundland, as inferred by tag recov- eries in a seabird colony. Can. J. Fish. Aquat. Sci. 45(3): 568-71. Montevecchi, W. A., and R. A. Myers. 1997. Centurial and decadal oceanographic influences on changes in northern gannet populations and diets in the north-eest Atlantic: implications for climate change. ICES Journal of Marine Science 54:608-14. Peterman, R. M., S. G. Marinone, K. A. Thomson, I. D. Jardine, R. N. Crittenden, P. H. Leblond, and C. J. Walters. 1994. Simulation ofjuvenile sockeye salmon iOncorhynchus nerka) migrations in the Strait of Georgia, British Colum- bia. Fish. Oceanogr. 3l4):221-35. Reddin, D. G., and K. D. Friedland. 1993. Marine environmental factors influencing the move- ment and survival of Atlantic salmon. In D. Mills (ed.l, salmon in the sea, p. 79-193. Fishing News Books. London. Reddin, D. G., and W. M. Shearer. 1987. Sea-surface temperature and distribution of Atlan- tic salmon in the Northwest Atlantic Ocean. Am. Fish. Soc. Synip. Ser. 1:262-75. Reddin, D. G., and P. B. Short. 1991. Postsmolt Atlantic salmon (Salmo salar) in the La- brador Sea. Can. J. Fish. Aquat. Sci. 48( 1 >:2-6. Robitaille, J. A., Y. Cote, G. Shooner, and G. Hayeur. 1986. Growth and maturation patterns of Atlantic salmon, Salmo salar, in the Koksoak River, LIngava, Quebec. In D. J. Meerburg (ed.), Salmonid age at maturity, p. 62- 69. Can. Spec. Publ. Fish. Aquat. Sci. 89. Shelton, R. G. J., W. R. Turrrcll, A. MacDonald, I. S. McLaren, and N. T. NocoH. 1997. Records of postsmolt Atlantic salmon, Salmo salar L., in the Faroe-Shetland Channel in June 1996. Fisher- ies Research (Amsterdam) 31(l-2):159-62. Thorpe, J. 1994. Salmonid fishes and the estuarine environment. Aquaculture 17( lA):76-93. 482 Abstract.— Using improved indices of age-1 recruitment and parent stock. I investigated the existence of a stock- recruitment relationship for ocean shrimp, Pandalusjordani. I also exam- ined the effects of large catches of egg- bearing shrimp in April on subsequent recruitment. After known environmen- tal influences were accounted for, re- gression analysis revealed evidence of a quasilinear relationship between re- cruitment and either spawning stock or egg production. The stock-recruitment relationship showed that average re- cruitment increases if higher spawner abundance is maintained in years of low abundance. The recruitment-stock relationship, however, showed that low abundance of spawning fish was mostly a result of environmentally driven re- cruitment failures, rather than just fishery impacts. Fishing did contribute to the low numbers of spawning fish in 1993, and catches of egg-bearing shrimp in April 1989 may have also de- pressed the 1990 year class. The strong recruitment-stock relationship for ocean shrimp suggests that an increase in escapement of all ages of shrimp in response to a year-class failure may do more to bolster recruitment than tra- ditional strategies aimed at px-otecting age-1 shrimp. A new method for indexing spawning stock and recruitment in ocean shrimp, Pandalus jordani, and preliminary evidence for a stock-recruitment relationship Robert W. Hannah Oregon Department of Fish and Wildlife Marine Program 2040 S E. Marine Science Drive Newport, Oregon 97365 E-mail address bobhannatimhrnscorstedu Manuscript accepted llAugust 1998. Fish. Bull. 97:482-494 (1999). Recruitment of the ocean shrimp, Pandalus jordani , to the fishery in waters off Oregon has been shown to be influenced strongly by ocean conditions during early larval stages (Hannah, 1993). Specifically, sea level height (SLH) during the month of April, shortly after the bulk of larval release, is strongly negatively correlated with recruit- ment of age-1 shrimp to the fishery the following year ( Hannah, 1993 ). Although the mechanism linking recruitment to April SLH remains unclear, April SLH probably inte- grates the strength and timing of the spring transition in coastal cur- rents (Huyer et al., 1979). Ocean shrimp larvae inhabit the near- surface waters shortly after release, occupying progressively deeper strata as they develop (Rothlisberg, 1975; Rothlisberg and Miller, 1983). The nature of the spring transition probably influences the alongshore transport of early larvae, as well as the nearshore temperature regime, either of which could strongly influ- ence shrimp larval survival (Rothlis- berg, 1975; Rothlisberg and Miller, 1983; Hannah, 1993; McConnaughey etal., 1994). Attempts to relate ocean shrimp recruitment to the size of the par- ent spawning stock, even allowing for the influence of April SLH, have been unsuccessful (Abramson and Tomlinson, 1972; Gotshall, 1972; Geibel and Heimann, 1976; Hannah, 1993 ). Efforts to date have, to a large degree, relied on assumptions of constant catchability and natural mortality. Recently, both natural mortality rates and the catchability coefficient have been shown to be quite variable for ocean shrimp (Hannah, 1995). I have suggested previously that variation in catch- ability and natural mortality may have created large enough errors in prior indices of shrimp stock and recruitment as to obscure an under- lying relationship between these two variables (Hannah, 1995). In the present study, new indices of ocean shrimp recruitment and spawning stock were developed which do not rely, or rely only very minimally, on assumptions of con- stant catchability and natural mor- tality. The main objective of this study was to re-examine the rela- tionship between shrimp recruit- ment and both shrimp spawning stock and April SLH in the year of larval release, using these improved indices. Another problem with earlier studies of ocean shrimp stock size and recruitment is that they have relied on measures of parent stock or spawning biomass, rather than on more direct measures of repro- ductive output, such as estimates of Hannah: A new method for indexing spawning stock and recruitment for Pandalus jordani 483 egg production (Rothschild and Fogarty, 1989; Hannah et al., 1995). A second objective of this study was to construct, in addition to an improved index of spawning stock, an index of egg production, and evaluate the relationship between recruitment and egg production. One of several challenges in con- structing such an index is that the trawl fishery har- vests egg-bearing female shrimp in late October, just before the fishing season ends, and again in early April, as fishing resumes. It can be argued that be- cause of reduced fishing effort late in the season in most years and because of the high natural mortal- ity rates experienced by ocean shrimp over the win- ter, the harvest of egg-bearing shrimp in October should have a negligible impact on total lar\'al re- lease the next spring. However, in April, when lar- val release is imminent for any shrimp still bearing eggs, the impact of the fishery could be more detri- mental. Also, with much of the larval release taking place before the fishing season opens on 1 April, fish- ery impacts are concentrated on late-release larvae. If, as postulated in the prior study (Hannah, 1993), the timing of the spring transition is critical to lar- val survival or retention in the study area, the fish- ery may be impacting the larvae with the greatest chances for success in a poor environment: those des- tined for release after a late spring transition. A third objective of this study was to estimate the harvest of egg-bearing shrimp in April and examine the impact that April harvest of egg-bearing shrimp may have had on subsequent recruitment. Materials and methods In my earlier study of ocean shrimp stock and re- cruitment (Hannah, 1993). I used an age-structured index of recruitment calculated by summing two com- ponents. The first component was an estimate of the age-2 shrimp population in April (year t), calculated from the average fishery catch per unit of effort (CPUE) of age-2 shrimp m April and May, divided by an assumed catchability coefficient calculated from data in Geibel and Heimann ( 1976). The second com- ponent was simply a sum of the catch of age- 1 shrimp during the prior year it-l ), with each monthly catch discounted to 1 April of year t by using an assumed monthly natural mortality rate of 0.096 (Gotshall, 1972). In essence, this index was a crude estimate of what the age-2 shrimp population would have been in the absence of fishing, on 1 April of the age-2 year (Hannah, 1993). To index the shrimp spawning stock in the earlier study, I used a simple average of Sep- tember-October shrimp fishery CPUE. Both indices depended heavily on an assumption of constant catchability. The recruitment index also relied on an assumption of constant natural mortality. The relationship between CPUE, population size, geographic stock area, and the catchability coefficient can be described by the equations CPUE = paNA-^ q = paA~^, (1) (2) where p = the proportion of shrimp within the sweep of the gear that will be captured (the el- emental efficiency); A = the geographic stock area (stock area); a = the area covered by a single sweep of the gear; b - a coefficient that describes how q will vary with stock area; N= the population size; and q = the catchability coefficient (Winters and Wheeler, 1985). Accordingly, an assumption of constant catchability is equivalent to assuming that stock area is constant. However, stock area for ocean shrimp has been shown to be roughly proportional to stock abundance and to vary substantially between years (Hannah, 1995; Hannah, 1997). In the present study, I incorporated stock area estimates into both the shrimp recruit- ment and spawner indices, making an assumption of constant catchability unnecessary. I did assume that 6 in the above equation was equal to 1, an as- sumption that has been previously shown to be rea- sonable, according to a roughly linear relationship between stock area and abundance (Hannah, 1995). My approach also assumes that the elemental effi- ciency of shrimp trawl gear, p, is constant. Because the assumed rate of natural mortality was used in the prior study (Hannah, 1993) only to dis- count age-1 catches forward in time to 1 April of the age-2 year, the simplest way to reduce the influence of variable natural mortality on the recruitment in- dex is to index the population at an earlier age. This is a useful approach; however, age-1 shrimp must still be fully recruited to the fishery at the time cho- sen. Using data from the early years of the shrimp fishery, Lo ( 1978) showed that age-1 shrimp were in- completely recruited to the California shrimp fish- ery, especially early in the April-October season. A variety of data suggest that, since about 1979, age-1 shrimp have been fully recruited to the trawl fishery by June or July of each year Prior to 1974, Oregon required a minimum codend mesh of 34.9-mm ( 1 3/8 in) stretch measure. Since this date, minimum codend mesh size has been unregulated in the Or- 484 Fishery Bulletin 97(3), 1999 egon fishery, although CaHfornia still requires a mini- mum codend mesh of 34.9 mm. Prior to 1978, shrimp growth was slower (Hannah and Jones, 1991) and age-1 shrimp in the study area (Fig. 1) were clearly not fully recruited to trawl gear until very late in the season (ODFW^ ). After 1978, shrimp growth in- creased, possibly owing to reduced density from heavy fishing, although other factors were also in- volved, notably accelerated sex change that may have been a response to reduced numbers of older shrimp (Charnovetal., 1978; Hannah and Jones, 1991 1. Av- erage codend mesh size in the Oregon fieet also de- creased sometime during the 1970s, averaging 30.2 mm by 1981, decreasing to 28.7 mm by 1991-92 (Jones et al., 1996 ). Accordingly, in this study I chose the August-September time period to index the age-1 shrimp population, and restricted my analysis to the years after 1979. With the months of August-Septem- ber, the population was indexed eight months earlier than in the previous study, but at a time period late enough that age-1 shrimp could reasonably be assumed to be fully recruited to the trawl gear being fished. To index shrimp recruitment in the present study, I used the average August-September CPUE for age-1 shrimp only as an index of density and multi- plied this index by stock area (ha) for the same year class (Hannah, 1997). In this case, CPUE was ex- pressed as shrimp per hectare trawled, using an av- erage estimate of 5.93 ha trawled per single-rig equivalent hour fished (Hannah, 1995). To account for years in which early season catches of age-1 shrimp were high, I added in the catch of age-1 shrimp for the months of April-July of the same year. Thus, recruitment was calculated as R, = iD,A^) + C, (3) where R D recruitment of age-1 shrimp in year /; the average fishei-y CPUE of age-1 shrimp in August and September of year t\ A = the stock area for the age-1 year class in year t\ and C = the summed fishery catch of age-1 shrimp in the months April to July of year t. Use of these additional catches for April-July does rely on an assumption of constant natural mortality, although the importance of this assumption is clearly reduced. Age-1 shrimp CPUE for Pacific States Ma- rine Fisheries Commission (PSMFC) statistical ar- eas 82-88 was obtained from Zirges et al. ( 1982) and ' ODFW (Oregon Department of Fish and Wildlife i. present. 2040 SE Marine Science Dr., Newport, OR. data. Stock Unit Washington Oregon Ocean shrimp stock area - 1984 Ocean shrimp stock area - 1988 California 1981 to Unpubl. Figure 1 Location of commercial concentrations of ocean shrimp (Pandalus jordani) off coastal Oregon, in Pacific States Marine Fisheries Commission statistical areas 82-88. Dark areas show the approxi- mate minimum areal extent of the shrimp grounds (1984) and the lighter shaded areas show the largest areal extent observed from 1980 to 96 1 1988). Hannah et al. ( 1997). The collection and analysis of biological samples from the commercial catch has been described by Hannah and Jones ( 1991 ). The data used to construct the recruitment and spawning stock indices developed in this study dif- fer in some respects from the data used for the prior recruitment study. In the prior study, logbook data were available from Washington, Oregon, and Cali- fornia, such that a complete accounting of catch and effort was possible for PSMFC areas 82-92 (Fig. 1). Beginning in 1992, logbook data from the states of California and Washington became unavailable. Ac- cordingly, the data used in this study was based on Oregon landings for the entire study period. To mini- mize the impact of the missing information on the indices developed, the study area was limited to ar- eas 82-88 (Fig. 1). I believe that the missing infor- mation will create very minimal error in the new indices for several reasons. First, Oregon landings comprise the great majority of the removals of shrimp from statistical areas 82-88. The average percent- Hannah: A new method for indexing spawning stock and recruitment for Pandalus jordanl 485 age of total catch from these areas landed in Oregon from 1980 to 1992 was 92.57f (Hannah et al., 1997). Second, the components of the new indices, with one exception, did not depend on a complete accounting of catch and effort from the study area but simply on a good estimate of average CPUE at age and accu- rate estimates of stock area. The one component of the recruitment index that was somewhat influenced by the missing catch data was the added age-1 shrimp catch from the months of April-July. A correction for this potential source of error is not available; how- ever, I believe the magnitude of error introduced is likely to be small in relation to the interannual varia- tion in the index itself.The use of CPUE as a direct index of density is equivalent to assuming that el- emental efficiency, p in Equation 2 above, equals 100%. This is clearly not reasonable, however; se- lecting a value for p is problematic. The principal importance of selecting a value for p is to obtain a proper scaling between the two components of the recruitment index: one based on CPUE and stock area; the other based on the early-season catch of age-1 shrimp. Proper scaling of these two components is less critical than it seems at first examination be- cause the two index components are actually closely correlated (/■=0.774 atp=0.5). Accordingly, a change in the scaling between the two components is ex- pected to have little influence on the pattern of time series variation in the resulting index. As a check on this assumption, all indices using CPUE as a den- sity index in this study were calculated with a range of p-values, and the results were then examined to see if they were sensitive to assumptions about p. For the primary analysis, I chose a p-value of 0.5 and for the sensitivity analysis I used values of 0.25 and 0.75. forp, following my earlier examination of shrimp mortality rates (Hannah, 1995). Commercial logbook data were used to generate estimates of stock area for each shrimp year class. The methods used to estimate stock area, including the correction of estimates for variation in sampling rates, are described in detail in Hannah ( 1997 ). Stock area was estimated by using logbook data from June of the recruit year through May of the following year, the time period during v/hich age-1 shrimp contrib- ute most heavily and reliably to the catch. For a few of the years in the 1980-96 time series, the fishery failed to target age-1 shrimp adequately, and har- vested primarily the remaining age-2 and older shrimp from prior year classes. For those years, stock area was estimated from a linear regression of stock area on a simple virtual population estimate for that year class, as described in Hannah (1997). To index the ocean shrimp spawning stock, I used an approach similar to that used for the recruitment index. For each spawning year, I used the average age-specific CPUE in September-October as an in- dex of age-specific density. Once again, these esti- mates of density were expanded by using assumed levels of elemental trawl efficiency ranging from 0.25 to 0.75, as discussed above. I then multiplied each density index by the stock-area estimate for that year class, as calculated in the year of recruitment. In using this approach I assumed that the geographic distribution of each newly recruited year class was established at settlement and was persistent, and that local shrimp density was modified by fishing and natural mortality. This assumption is supported by several observations, including the obvious auto- correlation in stock area estimates noted previously (Hannah, 1995), the lack of any evidence for migra- tion in this species and the approximately linear rela- tionship between stock area and abundance. There is some evidence fi'om sea bed drifter recoveries ( ODFW^ ) that suggests slow gyres in bottom currents may help retain or concentrate shrimp in some of the major shrimp beds. However, it is unknown whether this ef- fect is sufficiently large to actually alter the areal extent of a shrimp year class after settlement. An egg production index for ocean shrimp was cal- culated by expanding the spawner index with avail- able shrimp biological data (ODFWM. The biological data used included the mean percentage of females and mean female shrimp carapace length, by age, from samples of the fishery in October. For a few years, October samples were unavailable and Sep- tember samples were used. I used a pooled length- fecundity relationship from Hannah et al. ( 1995) to estimate mean fecundity for each age group of fe- male shrimp based on mean carapace length data. The April harvest of egg-bearing females was esti- mated for the years 1979-95 in a straightforward manner. April shrimp catch, expressed as numbers of shrimp, for each PSMFC statistical area, was ob- tained from Hannah et al. ( 1997). Catch in numbers was multiplied by the average percentage of egg-bear- ing shrimp estimated for each PSMFC area from bio- logical samples of the April catch (ODFWM. The catch of egg-bearing shrimp from each of the four areas (Fig. 1) was then summed. Because the percentage of egg-bearing shrimp de- clined throughout April, a time-stratified approach was used to estimate the average percentage of egg- bearing shrimp. One problem with this approach is variation in sample coverage. Although the total number of shrimp examined from the April catch has been fairly consistent, averaging about 2900 shrimp from 1980 to 1996, the early and late portions of April have not always been sampled evenly. For example, some years that had excellent sample coverage in 486 Fishery Bulletin 97(3), 1999 10^ 7.5 01 < 5- 2.5 the first two weeks of April had no samples at all for the second half of the month. To make maximum use of the data available, estimates of the percentage of egg-bearing shrimp from individual samples were averaged for the first and second halves of the month, respectively. These semi- monthly figures were then averaged to produce an overall mean percent- age of egg-bearing shrimp in the catch for each April. When samples were miss- ing from the latter half of the month, a zero level of ovigerous shrimp was as- sumed. When samples were missing for the first half of the month, the level of ovigerous shrimp observed for the sec- ond half was assumed as a minimum estimate for the first half of April. Ac- cordingly, the estimates presented in the present study are minimum estimates of the catch of egg-bearing shrimp. To determine when, during the month of April, egg-bearing shrimp had declined to a minimal component of the catch, I constructed a scatter graph of the raw percentages of egg-bearing females by date, using existing data for 1961-97 (ODFWM. Area 82 (Fig. 1) was excluded from this analysis because the incidence of egg-bearing shrimp in this area was generally very low in April. A stepwise process was used to evaluate the rela- tionships between the new recruitment, spawner, and egg production indices, calculated at assumed val- ues of 0.25., 0.50 and 0.75 for trawl efficiency. Fol- lowing the findings of my earlier study (Hannah, 1993), the first step in this analysis was to regress log (natural log unless noted) recruitment against April SLH at Crescent City, California, to verify that SLH still explained a significant amount of the interannual variation in recruitment. Simple linear regression was used for this test. These SLH data were obtained from the National Oceanic and Atmo- spheric Administration's Ocean and Lake Levels Di- vision for the years 1979-95, corresponding to age-1 catch years 1980-96. Next, the residuals were ex- amined and compared graphically and by means of linear regression to the spawner and egg production indices, and outliers were examined in light of the estimates of the catch of egg-bearing shrimp in April. Finally, some multivariate models incorporating April SLH and the spawner and egg production indi- ces were fitted by using multiple regression. Garcia ( 1983) and others have stressed that many of the stock-recruitment relationships that have been demonstrated for shrimp stocks are really statisti- 80 60 40 -20 1980 1985 1990 1995 Year Figure 2 Time series of the recruitment index, calculated assuming elemental trawl efficiency of 0.50. and the fishery catch of egg-bearing shrimp from April of the year of larval release, by year of age-1 recruitment, 1980-96. cal artifacts caused by a strong recruitment-stock relationship (the reverse of a stock-recruitment re- lationship, rather spawning stock in year / as a func- tion of recruitment that same year) in combination with serially autocorrelated, environmentally driven, recruitment. To test for such a possibility in the present analysis of ocean shrimp recruitment, I re- gressed the new recruitment index against itself at a lag of one year. I also examined the recruitment- stock relationship further by graphically comparing the spawning index, broken out by age class, to the recruitment index from the same year. Results The recruitment index (Fig. 2; Table 1; all indices calculated assuming 0.50 for trawl efficiency, unless noted) shows wide variation in ocean shrimp recruit- ment, especially after 1986 (year of age-1 recruit- ment, unless noted). Recruitment in 1983, 1984, 1990, and 1993 was very low, whereas very high re- cruitment was observed for the 1987-89 and 1992 year classes. These four large year classes had a very significant influence on total landings from the shrimp fishery during the years included in this study. Although they represent less than one third of the years in the time series, they contributed roughly 60% of the total landings of shrimp. The log of the shrimp recruitment index was stongly negatively correlated with April SLH at Cres- cent City, California, in the year of larval release Hannah: A new method for indexing spawning stock and recruitment for Pandalus /ordani 487 Table 1 Ocean shrim p population data u sed for regression analyses , including April SLH, geograph c stock area (year is June -May), and recruitment spawner, and egg product on indices, by ca len dar year (no lags). Recruitment, spawner, and egg product ion indices are calculated assuming elemental trawl efficiency of 0 50. Recruitment April Geographic Spawning Egg index SLH stock area abundance production Year (millions — age 1) (cml (1000 ha 1 (millions) (billions) 1979 — 214.9 — — — 1980 3122.0 218.2 450.6 — — 1981 2455.0 210.9 477,2 — — 1982 2789.1 226.5 344.1 2133.0 2002.9 1983 368.6 229.2 172.9 604.0 717.1 1984 884.5 211.5 158,7 1150.7 1257.4 1985 2798.4 213.1 319,4 3049.7 1913,4 1986 3148.7 212.5 460,5 2293,1 1616,0 1987 6840.2 209.1 464.6 3760,3 2558.0 1988 9388.1 214.9 605.0 7332.6 5783.8 1989 7078.9 222.2 585.1 6872.6 4845.5 1990 429.2 219.2 208.8 1652,2 3583.3 1991 4312.5 212.5 478.1 2141.4 1590.4 1992 8729.9 230.7 406.3 4519.7 4146.1 1993 649.9 227.4 200.8 921,4 1670.2 1994 2608.4 215.0 364.1 1465.3 1291.0 1995 768.6 220.0 215.7 — — 1996 2862.8 — 362.0 — — (Fig. 3), The linear regression model (Table 2; model 1) was highly significant, with April SLH explaining 48*7?^ (unadjusted for degrees of freedom, unless noted) of the variation in log recruits. More importantly, the linear relationship is not based heavily on just a few points but appears quite general in nature, confirm- ing a routine relationship between recruitment and April SLH shortly after the peak of larval release. The residuals from the regression of log recruits on April SLH, graphed against the spawner index, show a pattern that is suggestive of a quasilinear stock-recruitment relationship, with the 1990 year class as an exceptional outlier (Fig. 4). The same re- siduals regressed against the egg production index show a similar pattern (Fig. 5). The time series of recruitment and the April catch of egg-bearing shrimp from the year of larval release (Fig. 2) shows that the year class released in 1989 ( 1990 year class) may have been heavily impacted by the trawl fish- ery, owing to an extremely high harvest of late egg- bearing shrimp. This occurred in combination with a high April sea level, suggestive of a late spring tran- sition (Table 1). Data on the percentage of ovigerous females in the April catch (Fig. 6) suggest that post- ponement of the opening of the season to April 20'^ each year could eliminate this impact. 23.0- • • • • y= -0,104x+ 44.159 /2 = 0 482 E (J 22.0- ♦ X* * o o 21.0- « \. ♦ z 20.0- 19.0- ' ' ' ' 1 ' ' ' ' 1 ' • 1 1 205 210 215 220 225 230 235 April sea level at Crescent City (cm) Figure 3 Linear regression of the natural log of the shrimp recruitment index, calculated assum- ing elemental trawl efficiency of 0.50, versus mean sea level height at Crescent City, Cali- fornia, during April of the year of larval release. When the 1990 year class was excluded, neither scatter graph (Figs, 4-5 ) exhibited sufficient evidence of curvature to indicate the most appropriate form 488 Fishery Bulletin 97(3), 1999 1.2- « 0, 0-8l o o ^ -g 0.4- Ui 2 g 0- 1 -0.4- ^ -0.8- o o o o o o o -1.2- o 0 2 4 6 8 Spawner abundance index (billions) Figure 4 Residuals from regression of Ini recruits) on April SLH at Crescent City. California, versus the ocean shrimp spawner index from the par- ent year. Both indices were calculated assum- ing elemental trawl efficiency of 0.50, 9 tr 1.5-, 1 0.5 0 -0.5 -1 -1.5 o o o ,oO 900 — ' 1 ' 1 ' 1 0 2,000 4,000 6,000 Egg production index (billions) Figure 5 Residuals from the regression of ln( recruits) on April mean sea level height at Crescent City. California, versus the ocean shrimp egg pro- duction index from the parent year Both indi- ces were calculated assuming elemental trawl efficiency of 0.50. of the stock-recruitment relationship for ocean shrimp. If the 1990 year class was included, the graphs could be interpreted as supporting a dome- shaped stock-recruitment curve, or could even be interpreted as evidence of no relationship between stock and recruitment at all. Given the extremely large fishery impact on egg-bearing females in the 30- 20- Area 88 K * ""i's •'"iiSJii X M X „ 10 20 60 ID O ^ 40 20- Area 86 X fgii'ihilgiiiiiiiii ,i8. 10 20 ID- Area 84 «« SwSx X^XxXB^K* «XX Date in April Figure 6 The percentage of egg-bearing shrimp in April samples of the ocean shrimp fishery, for Pacific States Marine Fisher- ies Commission areas 84-88. 1961-97. spring of 1989 (Fig 2), in combination with the fact that April has been shown to be a critical period for larval survival, excluding the 1990 data point seems the most conservative approach. With the 1990 year class excluded, a multivariate model incorporating April SLH and the log of the spawner index, was highly significant (P<0.001 1 and explained 79*^ of the variation in log recruitment (Table 2, model 2). The comparable multivariate model, incorporating the egg production index, was also highly significant (P<0.01) and explained 68^^ of the variation in log recruitment (Table 2, model 3). Both models incor- porated April SLH and log spawners or eggs with negative and positive slopes, respectively. A linear regression of the recruitment index on itself, at a lag of one year, was nonsignificant (P>0.05) and showed little evidence of serial autocorrelation, suggesting that the evidence presented in Figures 4 and 5, for a statistical dependence of recruitment on parent stock in ocean shrimp, may be valid. Hannah: A new method for indexing spawning stock and recruitment for Pandalus /ordani 489 Table 2 Results of univariate and multivariate regression analysis. Dependent variable is the natural log of the ocean shrimp recruit- ment index in year t. All population indices are calculated assuming elemental trawl efficiency of 0.50. Parameters and variables Coefficients 95% CI F-value i?2 (adjusted) P>F Model 1 Intercept 44.159 April SLH, , -0.104 ±0.059 Full Model 13.983 0.448 0.0020 Model 2 (1990 excluded) Intercept 29.108 April SLH,_, -0.112 ±0.047 29.388 0.0004 Ln (spawner index), j 0.796 ±0.502 12.893 0.0058 Full Model 16.550 0.739 0.0010 Model 3 (1990 excluded) Intercept 21.636 April SLH,_j -0.103 ±0,056 17.528 0.0024 Lnlegg index), ., 0.799 ±0.754 5.743 0.0401 Full Model 9.676 0.612 0.0057 Model 4 Intercept 27.531 April SLH,_, -0.119 ±0.053 26.037 0.0006 Ln(spawner index), .^ 0.957 ±0.628 11.879 0.0073 April catch of ovigerous Shrimp, 1 (millions) -0.035 ±0.022 13.006 0.0057 Full model 11.956 0.733 0.0017 A multivariate model, using the complete data se- ries and incorporating April SLH. the natural loga- rithm of the spawner index, and the April catch of egg-bearing shrimp from the year of larval release, was also highly significant (P<0.005) and explained 809c of the variation in log recruitment (Table 2, model 4). As expected, the coefficients for SLH and the catch of egg-bearing shrimp in April of the re- lease year were both negative, whereas the coeffi- cient for the natural logarithm of spawners was posi- tive. Fitting this three-variable model, with the 1990 year class excluded, resulted in the catch of egg-bear- ing shrimp not contributing significantly to the over- all model fit, suggesting that this variable does not have general predictive value for the recruitment time series but is useful only in explaining the very low recruitment in 1990. Despite efforts to de-emphasize the importance of age-1 catches in the recruitment series by indexing the stock at an earlier age, these early catches still represented a large component of the index (Fig. 7). This finding suggests that the assumption of con- stant natural mortality could still be inducing errors in the recruitment index. However, when the analy- ses discussed above were conducted on the recruit- ment index excluding the April-July age-1 catches, the results were very similar. This similarity argues that the findings of this study are not sensitive to failure of the assumption of constant natural mor- tality, as it is employed in calculating the recruit- ment index. The results were also insensitive to the assumed level of elemental trawl efficiency (Table 3, models 5 and 6). The data in Figure 4 suggest that if shrimp spawner abundance could be maintained above some threshold (about 1.3 billion shrimp for an assump- tion of 0.50 for elemental trawl efficiency), the abil- ity to produce large shrnnp year classes in favorable environmental conditions could be preserved in all years. A similar threshold could also be identified in terms of egg production (Fig. 5). Examination of the relationship between age-1 spawners, older spawn- ers, and the recruitment index from the same calen- dar year (Fig. 8 ), suggests a strong recruitment-stock relationship for ocean shrimp and a minor role for the fishery in determining spawner abundance. In 490 Fishery Bulletin 97(3). 1999 Table 3 Results of multivariate regression analysis using alternative assumed va lues for elemental trawl eff ciency. Dependent variable | for models 5 and 6 is the natural log of the ocean shrimp recruitment index in year t. calculated assuming elemental trawl efficiency of 0.25 and 0.75, respect vely. Parameters and variables Coefficients 95% CI F-value R^ (adjusted) P>F Model 5 efficiency of 0.25 Intercept 27.583 April SLH,_j -0.122 ±0.056 24.306 0.0008 Ln (spawner index), _, 0.976 ±0.666 11.022 0.0089 April catch of ovigerous Shrimp,_[ (millions! -0.038 ±0.024 13.252 0.0054 Full model 11.500 0.724 0.0020 Model 6 efficiency of 0.75 Intercept 27.437 April SLH,_i -0.117 ±0.051 26.829 0.0006 Ln (spawner index),., 0.946 ±0.608 12.378 0.0065 April catch of ovigerous -0.034 ±0.022 12.683 0.0061 Shrimp, , (millionsl Full model 12.107 0.735 0.0016 g 2 12. 0^ 10,0 219 mm FL) which were chosen as they approximate the age of the fish within each size class as determined from the von Bertalanffy growth equa- tion derived by Machias et al. ( 1998). Each month 20-30 specimens were preserved in lO'^ buffered formalin immediately after capture for stomach-content analysis. Samples were taken to the laboratory, measured to the nearest mm (FL), and weighed to the nearest 0.1 g. Thereafter, the stom- achs were removed and the contents wet weighed. Prey items were identified to the lowest possible taxo- nomic level, counted under a binocular microscope and weighed to the nearest 0.01 g. Prey species were measured to the nearest 0.1 mm by using an ocular micrometer or a vernier caliper, where possible. Fork length of the fish examined ranged from 41 to 185 mm, mean (±SD) FL was 87.2 (±26.9) mm. Data analysis In accordance with the procedure described by Carrothers (1980), the door spread of the trawl net was calculated for each haul and then multiplied by the speed of the boat and fishing time to estimate the total area sampled. The abundance offish caught was expressed as number of individuals per square nautical mile (nmi'-). making the comparison offish abundance between sampling stations possible. Af- ter logarithmic transformations (Middleton and Musick, 1986; Stefanescu et al., 1992), mean abun- dances (number/nmi') were calculated for each cruise: 1 ) for 20-m depth intervals, 2) for 50-m depth intervals, and 3 ) per zone ( I, II, III ), as well as for each 1°C temperature interval. Salinity showed little vaiia- tion (Table 1), having no effect on the distribution of fishes on the Cretan shelf (Tsimenides et al., 1991). Analysis of variance showed no significant differ- ences in the mean abundance among the three sur- veys in each season, nor among each zone during the same season (0.3770, indicating a positive allometry on full stomachs, whereas no differences on the estimated probabilities were detected between the two years (^test=-0.50, P>0.05). Composition of the diet There were at least 58 different prey species belong- ing to four major groups (decapods, small crusta- Ldbropoulou et aL: Habitat selection and diet of luvenile Pagrus pagrus 499 Table 2 Correlation analysis between Pagrus pagn/s coefficient, P: significant level of probability. abundance and fork length with depth and temperature, r: Pearson correlation Abundance Mean length fork (mm) Minimum fork length (mm) Maximum fork length (mm) r P r P r P r P Depth (m) Spring -0.524 0.03 0.084 0.79 0.207 0.52 0.217 0.50 Summer -0.621 0.00 0.693 0.00 0.670 0.00 0.689 0.00 Winter -0.598 0.00 0.701 0.00 0.685 0.01 0.167 0.58 Temperature ('O Spring 0.224 0.10 0.633 0.07 0.070 0.85 0.585 0.10 Summer 0.787 0.00 0.401 0.25 0.510 0.13 0.621 0.06 Winter 0.349 0.01 0.138 0.74 0.097 0.82 0.299 0.47 Feeding indices of Pag Table 3 ■us pagrus from th e Cretan shelf in relation to size. Size class Fork length (mm) No. of stomachs analysed No. of full stomachs Mean no. of prey items per stomach' Mean weight prey items (g) per stomach' No. of prey species Diet breadth (Levin's index)- Number Biomass 0 1 2 42-104 105-145 146-185 450 161 23 271 117 20 2.96° (± 0.23) 1.98'' (± 0.20) 2.23'' (± 0.30) 0.17" (± 0.01) 0.46'' (± 0.01) 0.77' (±0.01) 45 36 27 0.39 (0.22-0.50) 0.38 (0.22-0.50) 0.373 (0.21-0.49) 0.45 (0.31-0.56) 0.29 (0.15-0.38 0.20 (0.12-0.33) ' Standard error given in parentheses, different letters indicate significant differences among - Ranges in parentheses: 959c bootstrap confidence intervals means (Tukey's test). ceans, polychaetes, and fish) (Table 4). Polychaetes predominated in terms of percentage by number (35.1%), whereas decapods made up 54.2% of the to- tal weight of stomach contents. Small crustaceans were also consumed in fairly large proportions by number (25.2% ), but their contribution by weight was minor. Fish were a considerable dietary staple by weight ( 23.5% ), although their contribution by num- ber (6.7%) was comparatively low. At the species level, the thalassinid Upogebia tipica, the caprellid Phtisica marina, and the sedentary polychaete Terebellides stroemi were the most exploited prey. Among fish prey, Gobius niger was the species having greatest contribu- tion in the diet of red porgy. Diet breadth was found to be 0.39 (±0.01) for the numerical abundance and 0.31 (±0.07) for the biomass of the prey species. Food in relation to Fish size Although the contribution of prey groups, in terms of both number and weight, varied with fish size (Table 4), there were significant differences only be- tween ingestion of small crustaceans (^'^=28. 46, P<0.001), U. tipica (x^=15.38, F<0.001), and Liocarcinus maculatus (X'=8.24, P<0.01). Small crus- taceans occurred in the younger specimens (0 size class), whereas U. tipica and L. maculatus occurred in greater percentages in the diet offish correspond- ing to size classes 1 and 2. The total amount of food ingested varied significantly among size classes (P=51.95, P<0.001). Pair-wise group comparisons showed three homogeneous groups; the mean con- sumption rate per individual (i.e. g food/size class) 500 Fishery Bulletin 97(3), 1999 Spring H 1 1 1 h- Summer Winter 11 i ' M ' -M — I — \ — ^ — t-"^ ^ — (N — < — CS ^ — CN Depth (m) 3.5 ,' 3 I ^ 2.5 • ^ M M 1.5 - I 0.5 Spring Summer Winter ^i , ^ F^ ML _ML Depth (m) 3.5 3 2.5 Spring Summer Winter 2 1.5 TTl 1 0.5 0 L_J 1 1 1 1 ill , 1^:^ , — 1 — likid — 1 t-^^'.-^t 1 Zone I Zone II Zone III Zone I Zone II Depth Zone III Zone I Zone II Zone III Figure 2 Seasonal distribution of Pagrus pagrus abundance in different depth range.s on the Cretan continental shelf. Log(A'+ll = the average logarithm of the number of individuals per nmi-. increased with increasing size, with 0-size-class fish to exhibit the lowest consumption rate (Table 3). The mean number of prey items consumed decreased sig- nificantly with size {F=3.33, P<0.05) (Table 3). Breadth of diet decreased with fish size, both nu- merically (abundance) and with bidmass of prey species (Table 3). In general, values for dietary overlap between size classes were similar irrespective of the resource matrix used. A significantly high overlap was observed only between specimens of 1 and 2 size classes (Table 5). Seasonal variation in diet Diet composition was fairly consistent over the months (Fig. 5, A and B). Decapods were the most important prey throughout the year, especially dur- ing winter months and March, owing to the increase of the importance of the thalassinid U. tipica. Poly- chaetes were also an important component in the diet of red porgy throughout the year. Dietary breadth varied little, except for low values in winter months Labropoulou et al : Habitat selection and diet of juvenile Pagrus pagrus 501 3.5 3 2.5 2 1.5 1 0.5 0 20-40m SPRINCi 40-60m 60-80m >80m Jii n 1 f-i^^^H 1 ^ Jl_ m,m 01234012340123401234 Size classes SUMMER 3.5 3 ~. 2.5 + 2 Z M 1.5 o ^ 1 0.5 0 20-40m n 40-60m 60-80m >80m H i h I ML -i r ja. i — \ — I — I — i — f aJL 01234012340123401234 Size classes WINTER 3.5 3 2.5 - 2 1.5 1 0.5 0 20-40m 40-60m 60-80m >80m -I 1 H -Mh ,.iUni h' I I ' 'I I ■ 1 1 1 H 01234012340123401234 Size classes Figure 3 Seasonal distribution of the abundance of Pagrus pagrus size classes from the Cretan conti nental shelf LoglA'+ll = the average logarithm of number of individuals per nmi-. and March, when U. tipica dominated (Fig. 6, A and B). Relationship of length of predator to length of prey Significant positive correlations were found between the size of red porgy and the mean size of prey con- sumed (P<0.001, r-=0.85, n = 169). Furthermore, analysis of variance on the mean prey sizes consumed by the three size classes revealed significant differ- ences (F=93.7, P<0.001). The 0-year fish exploited prey with smallest mean size (18 ±6.6 mm), 1-year fish consumed prey 26.2 ±5.4 mm, and 2-year fish fed on the largest prey (41.6 ±5.8 mm). Discussion Juvenile red porgy on the Cretan continental shelf tend to occur in shallows (20-50 m), where the sub- 502 Fishery Bulletin 97(3), 1999 4.5 4 -3.5 3 + so o J 2.5 2 1.5 Spring Summer Winter 1 1 -r ""t ^ Temperature "C Figure 4 Seasonal distribution of Pagrus pagrus abudance in 1 'C temperature intervals, on the Cretan continental shelf. Log(A''+l ) = the average logarithm of number of individuals per nmi-. strate is sandy and interspersed with patches of al- gae and seagrass. At this depth range, bottom water temperature is the highest during all seasons. Thus, this depth preference is consistent with the tendency for red porgy to select warm temperatures in rela- tion to those available in other areas of its distribu- tion (Manooch and Huntsman, 1977; Manooch and Hassler, 1978; Pajuelo and Lorenzo, 1996). Although red porgy were found to occur down to 250 m depth, their densities were significantly greater at shallow stations. Furthermore, specimens smaller than 186 mm predominated in the trawl catches, whereas rela- tive abundances of larger specimens was low. High densities of red porgy in shallow waters were due to the large number of juveniles, i.e. specimens that have not reached sexual maturity (Manooch and Hassler. 1978). The lack of larger fish from the trawl catches in shallower depths could not be due to gear selectivity because larger specimens were occasion- ally found in deeper waters, indicating that they could have also been caught in shallower depths, if present. A similar distribution pattern has also been reported for the Indo-Pacific species Pagrus auratus which has been studied in many different areas ( Azeta et al., 1980; Kingett and Choat, 1981; Tanaka, 1985). Size (or age) tends to be positively correlated with depth in many demersal fishes. Tremblay and Sinclair (19851 reported that the mean depth of oc- cuJTence tended to increase with age of cod in the southern Gulf of St. Lawrence. Similarly, Sinclair (1992) noted a positive correlation between age and median depth of cod on the eastern Scotian Shelf. Macpherson and Duarte( 1991) found that mean fish length increased with depth in most south-east At- lantic and north-west Mediterranean demersal fishes. Bathymetric trends in demersal fish distri- bution may be linked to other physical factors corre- lated with depth (e.g. temperature, salinity, bottom type). Temperature is a key factor in the metabolism of fishes and many studies have reported relation- ships between temperature and fish distribution ( Fry, 1971; Nakken and Raknes, 1987; Rose and Leggett, 1989; Macpherson and Duarte, 1991). However it is unlikely that fish distributions are determined by physical factors alone. Biotic factors, in particular prey abundance, have also been shown to be impor- tant determinants of distribution. Productivity gen- erally decreases with increasing depth and distance from land; therefore prey resources are greater in shallow waters. Haedrich and Rowe ( 1977) concluded that selection for mobility and metabolic efficiency should favor larger size in the relative barren deep sea, whereas Macpherson and Duarte (1991) sug- gested that the positive size-depth relationship in demersal fish may result from age-specific differences in temperature preferences. They argued that younger fish occupy warmer waters, where food supply and growth rates may be gi-eater, whereas older fish oc- cupy colder waters, where they may benefit from lower metabolic costs and greater longevity. The most striking aspect of our results, was the lack of variation in the bathymetric pattern of red porgy and the fact that depth distribution varied little among size classes, during all seasons. The scarcity of specimens larger than 187 mm in the bathymetric range >50 m is noteworthy, especially for a species with a life span more than 12 years (Vassilopoulou and Papaconstantinou, 1992). Thus the absence of mature specimens in trawl nets during the pre.sent Labropoulou et al : Habitat selection and diet of juvenile Pagrus pagrus 503 Table 4 Percentage contribution by numbe ■in) and by weight a') of the major prey taxa and species in the diet of each size class of Paerus pagrus from the Cretan shelf. Prej occurring in >1% of the total are given. (+ = 200 mm FL the importance of decapods, fish, and polychaetes declined rapidly with size, being replaced by anthozoans, brachyurans, and echinoderms. Unlike the results of the present study, Labropoulou et al : Habitat selection and (diet of juvenile Pagrus pagrus 505 A Numerical abundance 0.9 ^ 0.8 0.7 0.6 0.5 0.4 0.3 0.2 0.1 0 0.9 0.8 0.7 + 0.6 0.5 0.4 - 0.3 0.2 0.1 0 f \ Aug Sep Oct Nov Dec Feb Mar Apr May Jun .lul B Biomass H h Aug Sep Oct Nov Dec Feb Mar Apr May Jun Jul Figure 6 Monthly variations of Pagrus pagrus dietary breadth: numerical abundance (A): biomass (Bi. Error bars are 95'^r bootstrap confidence intervals. their results showed pronounced seasonal changes in the diet of red porgy. It is possible that these dif- ferences in feeding habits between juvenile and ma- ture individuals are associated with habitat change from soft bottom to hard substrates. Manooch ( 1977 ) stated that red porgy is an opportunistic browser feeding on a variety of invertebrates as well as small fish. He also noted that its diet appeared to be de- pendent on species availability, rather than prefer- ence or selection. These differences between our re- sults and those of the previous studies may be due in part to the fact that in both studies dietary analy- sis was based on mature specimens, probably reflect- ing the distinct spatial zones occupied by juvenile and mature specimens. According to our results, habitat of juvenile red porgy appears to be well separated from that of adult specimens. The absence of mature individuals in the trawlable fishing grounds indicates that recruitment in shallow waters is followed by an ontogenetic move- ment to a new habitat, at the time they reach sexual maturity. Further examination of this phenomenon is important from a biological, as well as from a man- agement, point of view. Red porgy have previously been described as a food generalist obtaining most of its food from the benthos and epibenthos as well Table 5 Diet overlap for the three size classes (0. 1, and 2)o{Pagn/s pagrus based on the numerical abundance and biomass of prey species. Ranges in parentheses: 95'7f bootstrap confi- dence intervals. Significantly high overlap values are in bold. Size class Biomass 0 1 1 2 0.44 (0.32-0.56) 0.31 (0.18-0.48) 0.73 (0.67-0.82) Size class Numerical abundance 0 1 1 2 0.51 (0.31-0.66) 0.36 (0.19-0.47) 0.72 (0.57-0.84) as a habitat generalist occupying a variety of habi- tats (Manooch, 1977; Manooch and Hassler, 1978; Pajuelo and Lorenzo, 1996). Patterns observed in the 506 Fishery Bulletin 97(3), 1999 present study suggest that habitat segregation is of importance: juvenile specimens are differentiated from mature ones on the basis of the bottom type and prey selection and to a lesser extent on the basis of depth. Literature cited Azeta, M., R. Ikemoto, and M. Azuma. 1980. Diurnal activity rhythms in O-group red sea bream, Pagrus major. Bull. Seikai. Reg. Fish. Res. Lab. 54:279- 289. Berg, J. 1979. Discussion of methods of investigation the food of fishes with reference to a preliminary study of the prey of Gobiusculus flavencens. Mar. Biol. 50:263-273. Bowen, S. H. 1983. Quantitative description of the diet, /n L. A. Nielsen and D. L. Johnson ledsl. Fisheries techniques p. 325- 336. Am. Fish. Soc, Bethesda, MD. Caddy. J. F. 1993. Some future perspectives for assessment and manage- ment of Mediterranean fisheries. Sci. Mar 57:121-130. Carrothers, P. J. G. 1980. Estimation of trawl door spread from wing spread. J. Northwest Atl. Fish. Sci. 1:81-89 Divanach, P., M. Kentouri, G. Charalambakis, F. Pouget, and A. Sterioti. 1993. Comparison of growth performance of six Mediterra- nean fish species reared under intensive farming condi- tions in Crete (Greece), in raceways with the use of self feeders. In G. Barnabe and P. Kestemont (eds.). Produc- tion, environment and quality, p. 285-297. Special Pub- lication 18, Ghent, Belgium. Efron, B., and R. Tibshirani. 1986. Bootstrap methods for standard errors, confidence intervals and other measures of statistical accuracy. Stat. Sci. 1:54-77. Fange, R., and D. Grove. 1979. Digestion. In W. S. Hoar, D. J. Randall, and J. R. Brett (eds I, Fish physiology, p. 161-260. Academic Press, New York, NY. Fry, F. E. J. 1971. The effect of environmental factors on the physiol- ogy of fish. In W. S. Hoar, D. J. Randall and J. R. Brett (eds.). Fish physiology, vol. VI, p. 1-98. Academic Press, New York, NY." Gibson, R. N.. and I. A. Ezzi. 1987. Feeding relationships of a demersal fish assemblage on the west coast of Scotland. ,J. Fish Biol. 31:5.5-69. Grove, D., and C. Crawford. 1980. Correlation between digestion rate and feeding fre- quency in the stomachles.s teleost Blennius pholis L. J. Fish Biol. 16:23.5-247. Haedrich, R. L., and G. T. Rowe. 1977. Megafaunal biomass in the deep sea. Nature 269: 141-142. Hall, S. J., D. Raffaelli, D. J. Basford, M. R. Robertson, and R. Fryer. 1990. The feeding relationships of the larger fish species in a Scottish sea loch, J. Fish Biol. 37: 775-791. Harris, P. J., and J. C. McCovern. 1997. Changes in the life history of the red porgy, Pagrus pagrus. from the southeastern United States, 1972- 1994. Fish. Bull. 95:732-747. Hawkins, A. D., N. M. Soofiani, and G. W. Smith. 1985. Growth and feeding of juvenile cod (Gadus morhua L.). J. Cons. Int. Explor Mer 42:11-32. Hurlbert, S. H. 1978. The measurement of niche overlap and some relatives. Ecology 59: 67-77. Hyslop, E. J. 1980. Stomach contents analysis-a review of methods and their application. J. Fish Biol. 17:411-429. Keast, A. 1978. Trophic and spatial interrelationships in the fish spe- cies of an Ontario temperature lake. Env. Biol. Fish. 3:7- 31. Kingett, P. D., and J. H. Choat. 1981. Analysis of density and distribution patterns in Chry'sophrys auratus (Pisces: Sparidae) within a reef en- vironment: an experimental approach. Mar Ecol. Prog. Ser 5:283-290. Kislialioglu, M. and R. N. Gibson. 1976. Prey "handling time" and its importance in food se- lection by the 15-spined stickleback. Spinachia spinachia (L. ). .J. Exp. Mar Biol. Ecol. 25:151-158. Krebs, C. J. 1989. Ecological methodology. Harper and Row, New York, NY, 654 p. Langton, R. W. 1982. Diet overlap between Atlantic Cod. Gadus morhua. Silver Hake. Merluccius bilmeans and fifteen other North- west Atlantic finfish. Fish. Bull. 80:74.5-759. Liem, K. F. 1984. Functional versality, speciation, and niche overlap: are fishes different? In D. G. Meyers and J. R. Strickler (eds), Trophic interactions within aquatic ecosystems, p. 269-305. Westview Press, Boulder. L' Abee-Lund, J. H., A. Langeland, B. Jonsson, and O. Ugedal. 1993. Spatial segregation by age and size in Arctic charr: a trade-off between feeding possibility and risk of pre- dation. J. Anim. Ecol. 62:160-168. Machias, A., N. Tsimenides, L. Kokokiris, and P. Divanach. 1998. Ring formation on otoliths and scales of Pagrus pagrus: a comparative study. J. Fish Biol. 52: 350-361. MacPherson, E. 1981. Resource partitioning in a Mediterranean demersal fish community Mar. Ecol. Prog. Ser 4:183-193. MacPherson E., and C. M. Duarte. 1991. Bathymetric trends in demersal fish size: is there a general relationship? Mar Ecol. Prog. Ser 71:103-112. Manooch, C. S. 1976. Reproductive cycle, fecundity and sex ratios of the red porgy, Pagrus pagrus (Pisces: Sparidae) in North Carolina." Fish. Bull. 74:775-781. 1977. Foods of the red porgy, Pagrus pagrus Linnaeus (Pi- sces: Sparidae), from North Carolina and South Carolina. Bull. Mar Sci. 27:776-787. Manooch, C. S., and W. W. Hassler. 1978. Synopsis of biological data on the red porgy, Pagrus pagrus (Linnaeus). U.S. Dep. Commer. NOAATech. Rep. NMFSCirc. 412:1-19. Manooch, C. S., and G. R. Huntsman. 1977. Age, growth and mortality of the red porgy, Pagrus pagrus. Trans. Am. Fish. Soc. 106:26-33. Martin, N. V. 1970. Long-term effects of diet on the biology of the lake Labropoulou et aL; Habitat selection and diet of luvenile Pagrus pagrus 507 trout and the fishery in Lake Opeongo, Ontario. J. Fish. Res. Board Can. 27:125-146. McCormick, M. I. 1989. Spatio-temporal patterns in the abundance and popu- lation structure of a large temperate reef fish. Mar Ecol. Prog. Ser 53:215-225. Middleton R. W., and J. A. Musick. 1986. The abundance and distribution of the family Macrouridae (Pisces: Gadiformes) in the Norfork Canyon area. Fish. Bull. 84:35-62. Miller, R. J. 1979. Relationships between habitat and feeding mecha- nisms in fishes. In R. Stroud and H. Clepper (eds). Preda- tor-prey systems in fisheries management, p. 269- 280. Sport Fishing Inst. Wash., D.C. Mittelbach, G. G. 1981. Foraging efficiency and body size: a study of optimal diet and habitat use by bluegills. Ecology 62:1370-1386. Nakken, O., and A. Raknes. 1987. The distribution and growth of Northeast Arctic cod in relation to bottom temperatures in the Barents Sea, 1978-1984. Fish. Res. 5:243-252. Norton, S. F. 1991. Habitat use and community structure in an assem- blage of cottid fishes. Ecology 72:2181-2192. Osenberg, G. W., G. G. Mittelbach, and P. C. Wainwright. 1992. Two-stage life histories in fish: the interaction be- tween juvenile competition and adult performance. Ecology 73:255-267. Pajuelo, J. G., and J. M. Lorenzo. 1996. Life history of the red porgy Pagrus pagrus (Teleostei: Sparidael off the Canary islands, central east Atlantic. Fish. Res. 28:163-177. Papaconstantinou, C, and E. Caragitsou. 1989. Feeding interaction between two sympatric species Pagrus pagrus and Phycis phycis around Kastellorizo Is- land (Dodecanese, Greece). Fish. Res. 7:329-342. Robb, A. P., and J. R. G. Hislop. 1980. The food of five gadoid species during the pelagic 0- group phase in the northern North Sea. J. Fish Biol. 16:199-217. Rose, G. A., and W. C. Leggett. 1989. Interactive effects of geophysically-forced sea tem- peratures and prey abundance on mesoscale coastal dis- tributions of a marine predator, Atlantic cod iGadus morhua). Can. .1 Fish. Aquat. Sci. 46:1904-1913. Ross, S. T. 1986. Resource partitioning in fish assemblages: a review of field studies. Copeia 1986:352-388. Ruiz, G. M., A. H. Hines, and M. H. Posey. 1993. Shallow water as a refuge habitat for fish and crus- taceans in non-vegetated estuaries: an example from Chesapeake Bay Mar Ecol. Prog. Ser 99:1-16. Sale, P. F. 1979. Habitat partitioning and competition in fish commu- nities. In R. Stroud and H. Clepper (eds I, Predator-prey systems in fisheries management, p. 323-331. Sport Fish- ing Inst. Wash., D. C. Schluter, D. 1994. Experimental evidence that competition promotes divergence in adaptive radiation. Science 266:798-801. Sinclair, A. 1992. Fish distribution and partial recruitment: the case of eastern Scotian Shelf cod. J. Northwest Atl. Fish. Sci. 13:15-24. Smith, P. W., and L. M. Page. 1969. The food of spotted bass in streams of the Wabash River drainage. Trans. Am. Fish. Soc. 98:647-651. Stearns, S. C. 1992. The evolution of life histories. Oxford Science Publ., Oxford, 264 p. Stefanescu, C, J. Rucabado, and D. Lloris. 1992. Depth-size trends in western Mediterranean demer- sal deep-sea fishes. Mar Ecol. Prog. Ser 81:205-213. Stergiou, K. I., and D. A. Pollard. 1994. A spatial analysis of the commercial fisheries catches from the Greek Aegean sea. Fish. Res. 20:109-135. Swain, D. P. 1993. Age- and density-dependent bathymetric pattern of Atlantic cod (Gadus morhua) in the Southern Gulf of St. Lawrence. Can. J. Fish. Aquat. Sci. 50: 1255-1264. Tanaka, M. 1985. Factors affecting the inshore migration of pelagic lar\al and demersal red sea bream Pagrus major to a nurs- ery ground. Trans. Am, Fish. Soc. 114:471-477. Tremblay, M. J., and M. Sinclair. 1985. Gulf of St. Lawrence cod: age-specific geographic dis- tributions and environmental occurrences from 1971 to 1981. Can. Tech. Rep. Fish. Aquat. Sci. 1387, 43 p. Tsimenides, N., G. Tserpes, A. Machias, and A. Kallianiotis. 1991. Distribution of fishes on the Cretan shelf J. Fish Biol. 39:661-672. Vassilopoulou, V., and C. Papaconstantinou. 1992. Age growth and mortality of the red porgy, Pagrus pagrus. in the eastern Mediterranean sea (Dodecanese, Greece). Vie Milieu 42:51-55. Vaughan, D. S., G. R. Huntsman, C. S. Manooch, F. C. Rohde, and G. F Ulrich. 1992. Population characteristics of the red porgy. Pagrus pagrus. stock off the Carolinas. Bull. Mar Sci. 50:1-20. Wallace, R. K., Jr. 1981. An assessment of diet -overlap indexes. Trans. Am. Fish. Soc. 110:72-76. Werner, E. E. 1979. Niche partitioning by food size in fish communi- ties. In R. Stroud and H. Clepper (eds). Predator-prey systems in fisheries management, p. 311-322. Sport Fish- ing Inst. Wash., D.C. Werner, E. E., and J. F. Gilliam. 1984. The ontogenetic niche and species interactions in size- structured populations. Annu. Rev. Ecol. Syst. 15:393-425. Werner, E. E., J. F. Gilliam, F. Hall, and G. G. Mittelbach. 1983. An experimental test of the effects of predation risk on habitat use in fish. Ecology 64:1540-1548. Werner, E. E ,and D. J. Hall. 1988. Ontogenetic habitat shifts in bluegill: the foraging- predation risk trade-off. Ecology 69: 1352-1366. Biostatis- tical analysis, 2nd ed. Prentice-Hall, Englewood Cliffs, NJ, 718p. 508 Abstract.— Fish assemblages of near- shore hardbottom habitats of southeast Florida were quantified at three sites from April 1994 to June 1996. Random 2 X 15 m transects were visually censused within two replicate areas at each site. The hardbottom at one site was buried by a dredge project to widen a beach one year into the study. A total of 394 transects were sampled. Eighty-six taxa (77 identified to species I from 36 families were censused. Grunts (Hae- mulidae) were the most diverse family (11 species), followed by the wrasses (Labridae) and parrotfishes (Scaridae) with seven and six species, respectively. The most abundant species were sail- ors choice i Haemulon parra I, silver porgy iDiplodusargenteus). and cocoa damself- ish iStegastes rariabilisi with mean abundances (individuals/transect) of 4.5, 3.8, and 3.7, respectively. Early life stages (newly settled, early juvenile, and juvenile) represented over 80'7f of the individuals at all sites. Newly settled stages of over 20 species were observed in association with hard- bottom reef structure. Outside of la- goons, nearshore hardbottom areas are the primary natural structures in shal- low waters of mainland Florida's east coast and were estimated to have nurs- ery value for 34 species of fishes. After one year, burial of approximately five ha of hardbottom habitat at one site lowered the numbers of individuals and species by over 30x and lOx, respec- tively. Due to their early ontogenetic stage, many of these species may not be adapted for high mobility in re- sponse to habitat burial. Dredging ef- fects may be amplified by burial prior to and during spring and summer peri- ods of peak lan.'al recruitment. Nearshore hardbottom fishes of southeast Florida and effects of habitat burial caused by dredging Kenyon C. Lindeman Division of Marine Biology and Fisheries, Rosenstiel School of Marine and Atmospheric Science, University of Miami 4600 Rickenbacker Cswy, Miami, Florida 33149 E-mail address (for K C Lindeman) klindeman:g'rsmas miami edu David B. Snyder Continental Shelf Associates, Inc. 759 Parkway St Jupiter, Flonda 33477 Manuscript accepted 28 August 1998. Fish. Bull. 97:508-525 (1999). The southeast coast of mainland Florida is within a biogeographic transition zone of high marine biodiversity (Briggs, 1974; Gilmore, 1995). This region is also undergo- ing some of the most rapid human population growth of any area of the United States (Culliton et al, 1990). Due to the economic and recre- ational value of beaches, substan- tial marine dredging projects (up to 1.5 X 10^ m'' of fill/project) are commony used to widen beaches that are subject to erosion in the area (ACOE, 1996). Nearshore hardbottom habitats are the pri- mary natural reef structures of this region at depths of 0-4 m and are often buried or indirectly affected by these projects. To date, no quanti- tative studies of the fish fauna of these habitats or the effects of beach dredge-and-fill projects on near- shore fishes are available (NRC, 1995). Nearshore hardbottom habitats of this area are derived from accre- tionary ridges of coquina mollusks, sand, and shell marl which lithified parallel to ancient shorelines dur- ing Pleistocene interglacial periods (Duane and Meisburger, 1969; Hoffmeister, 1974). The habitat complexity of these limestone struc- tures has been expanded by colonies of tube-building polychaete worms (Kirtley and Tanner, 1968) and other invertebrate and macroalgal species (Goldberg, 1973; Nelson, 1989; Nelson and Demetriades, 1992). In southeast Florida, most nearshore hardbottom structures are within 200 m of the shore. These habitats are often centrally located between mid shelf reefs to the east and estuarine habitats within inlets to the west. Therefore, they may serve as settlement habitats for immigrating larvae or as interme- diate nursery habitats for juveniles emigrating out of inlets (Vare, 1991; Lindeman, 1997a). Nonetheless, most administrative reviews have concluded that the fish habitat value of nearshore hardbottom and the effects of dredge-based beach restoration projects are minimal (e.g. ACOE, 1996). This study quantifies nearshore hardbottom fish assemblages on the southeast coast of mainland Florida over a 27-month period. The effects of dredge-fill placement were also examined because the hardbottom habitat at one site was buried on account of a beach restoration project 12 months into the study. Three primary objectives were ex- amined. First, spatial and temporal attributes of fish assemblages at Lindeman and Snyder: Nearshore hardbottom fishes of southeast Florida 509 CORAL COVE upiter Inlet CARLIN PARK Figure 1 Primary study sites for fish surveys of nearshore hardbottom habitats at Jupiter. Florida (26=56'N. 80-04'W). three undisturbed hardbottom sites were character- ized. Second, abundances of different life stages were compared to assess the potential nursery value of nearshore hardbottom habitat. Third, effects of dredge burial on numbers of individuals and species were compared between a site subjected to burial and a control site. Methods Study areas Fish abundances were quantitatively surveyed on two nearshore hardbottom sites approximately 2 km north (Coral Cove) and 2 km south (Carlin Park) of Jupiter Inlet. Florida (26 = 56'N, 80"04'W) from April 1994 through June 1996 (Fig. 1). Sampling at both sites extended approximately 100 m offshore to a depth of 4 m. Nearshore hardbottom of similar depth and structure at Ocean Ridge, immediately south of the South Lake Worth Inlet (26"31'N, 80 02'W) was also surveyed for comparative purposes during the summer of 1995. Weathered limestone outcroppings were common between depths of 0 and 4 m at all sites. These struc- tures have a variety of names (e.g. Anastasia forma- tion outcroppings, coquina reefs, worm reefs) but are referred to by their most common name, "nearshore hardbottom," in the present study. In some areas, the hardbottom extended 1.75 m above the bottom and was highly convoluted. Shoreward portions of the hardbottom were exposed at low tide. Epibiota consisted of a variety of invertebrates and algae. The 510 Fishery Bulletin 97(3), 1999 most widespread encrusting organism was the reef- building sabellariid worm Phragmatopoma lapidosa (=P. caudata: Kirtley, 1994), often covering over 50"^ of the hardbottom at all sites. A beach restoration project occurred at Carlin Park in March and April of 1995. More than 350,000 m'^ of beach-compatible sediments were excavated by a cutter-head dredge from a site 0.8 km offshore and hydraulically pumped along 1.8 km of shoreline. Bulldozers extended the fill seaward to an estimated width of 60 m. An estimated total of 4.9 to 5.7 ha ( 12-14 acres) of nearshore hardbottom was buried.^ Visual surveys of fishes were conducted for 12 months before burial and 15 months after burial at both Carlin Park (the impact site) and Coral Cove (the control site). Little or no hardbottom was observed in fish surveys at Carlin Park after the project. Survey protocol During each site visit, three to five transects within two adjacent areas (=6-10 total transects/site) were censused. These 2 x 15 m transects were randomly located at depths ranging from one to four m. Transects were deployed along random compass headings at random distances between successive transects. Random number tables were used prior to site visits to determine the compass headings and distances between transects (based on numbers of fin kicks). All fishes observed within one m of each side of the transect line were identified and tallied by a snorkeler The survey zone extended from the bottom to the surface and 2 m in front of the observer. Ledges and sand-rock interfaces were examined for fishes. Rocks were not overturned. The proportion of hardbottom to sand was estimated within each transect. An estimated 35"^ of the area within all transects was sand. Surveys were conducted between 0900 and 1700 and avoided twilight periods. To com- pare fish abundances at hardbottom and sand ar- eas, identical transect methods were used at near- shore sand plains greater than 50 m from any hard- bottom structures. Monthly visual censusing occurred from April 1994 to June 1996 as permitted by nearshore visibility and sea state. Discharges of turbid water from Jupiter Inlet and wave resuspension of fine sediments some- times resulted in turbidity levels that precluded sam- pling. Samples were obtained for all months except October through January when waves and turbidity were typically prohibitive. ' Davis, P. 1998. Palm Beach County Department of Environ- mental Resources Management, 3.32.3 Belvedere Rd.. Bldg. .502, W. Palm Beach, FL 33406. Personal commun. In addition to total abundances, early life stages were also enumerated. Fork length was used for size estimation. Following Lindeman (1986; 1997a), life stages of grunts [Haemulon and Anisotremus) were recorded as follows: newly settled (<2 cm), early ju- venile (2-5 cm), juvenile (5-15 cm), and adult (>15 cm). For other families, the same newly settled size range (<2 cm) was used. The early juvenile designa- tion was used only for grunts because of the distinct morphological features of the 2-5 cm size range (Lindeman, 1986). Juvenile and adult stages were based on size and pigment patterns reported in the literature (e.g. Robins and Ray, 1986; Humann, 1994). Species identifications of the newly settled or juve- nile stages for certain taxa were limited by very simi- lar morphological features (e.g. scarids, kyphosids, gerreids, haemulids, clupeids >. Some early stage iden- tifications were therefore recorded only at the genus or family level. Collections of small schools of newly settled grunts were made with hand nets at both Jupiter sites in 1994 and 1995 to supplement field identifications. All collections were deposited at the Florida Museum of Natural History, University of Florida. Data analyses To address the first objective of our study, two comple- mentary multivariate methods were used to spatially and temporally characterize the assemblages at the three sites. The second objective was addressed by univariate testing of the hypothesis that abundances of different life stages would not differ significantly within sites. The third objective was examined with the hypothesis that numbers of individuals and spe- cies would not differ significantly between an impact site where allmost all the hardbottom was buried and a control site that was unaffected by the burial. In univariate analyses, data were standardized as the mean number of individuals per transect and as the mean number of species per transect. Samples were temporally unbalanced owing to the inability to visually sample during portions of the winter. Therefore, to examine the first objective, multivariate ordination and classification (cluster analysis) of a samples-by-taxa matrix for the entire, unpooled data set were used. These analyses were performed on a data set of 31 samples ( 16 from Coral Cove, 12 from Carlin Park, and 3 from Ocean Ridge) and 61 taxa. Each sample represented a site visit where 6 to 10 transects were censused. Samples from Carlin Park did not include postdredging site visits because these samples contained few or no fishes. The 61 taxa were those remaining from a total of 86 after eliminating taxa occurring only once across all Lindeman and Snyder: Nearshore hardbottom fishes of southeast Florida 51 samples. Within each sample, counts for individual taxa were averaged over all transects to provide val- ues for the matrix. These values were log-trans- formed [logjQ(7i-t-l)] to prevent abundant taxa from dominating the ordination or classification results. The transformed matrix was analyzed by corre- spondence analysis (CA), a method that employs a two-way weighted averaging algorithm to produce simultaneous ordination of sites and taxa (Gauch, 1982; Jongman et al., 1995). These analyses were performed with the program CANOCO (ter Braak, 1988). From the same log-transformed data matrix, normal (samples) and inverse (taxa) resemblance matrices were generated by using the Bray-Curtis dissimilarity index (Bray and Curtis, 1957). Normal and inverse resemblance matrices were clustered separately by the unweighted paired-group method of averaging (UPGMA) (Sneath and Sokal, 1973). All dissimilarity and cluster analyses were computed with NTSYS-pc software (Rohlf, 1997). To address the second project objective, numbers of life stages per transect were compared within each site. Data were analyzed by using a parametric one- way ANOVA when variances were homogeneous (Bartlett's test). A posteriori comparisons of differ- ences among means employed Tukey's HSD test. Variances of numbers of life stages of grunts per transect at the two Jupiter sites remained heteroge- neous after logji/n-t-l ) transformation and a Kruskall- Wallis non parametric, single classification ANOVA was used. Probability was calculated using the x~ approximation. Two-sample ^tests for unequal vari- ances were used to compare numbers of individuals at hardbottom and natural sand sites. Only hard- bottom samples from months when natural sand sites were sampled (March and April 1995) were used for these tests. In all statistical tests, differences were considered significant at P<0.05. The third objective, examining dredging effects at the impact site (Carlin Park) and the control site (Coral Cove), employed a BACIPS (before after con- trol impact paired series) design (Stewart-Oaten et al., 1986; Osenberg and Schmitt, 1996). This ap- proach compares differences in variables between sites over time before and after the impact. The dif- ferences in the paired series were examined by two- sample Ntests by using the mean number of both individuals and species as the variables (Stewart- Oaten, 1996). Results A total of 352 transects was sampled at the two Ju- piter sites: 204 at Carlin Park and 148 at Coral Cove. a 60 40 ^ 20 5 - 10 15 Abundances among sites Coral Cove Carlin Park Ocean Ridge Figure 2 Mean number of individuals and species per transect (with 95% confidence intervals i for all nearshore hardbottom sites. Only predredging data were used for Carlin Park site. Sample sizes: Coral Cove: 148 transects; Carlin Park: 112; Ocean Ridge: 36. At Carlin Park, 112 transects were sampled before burial and 92 after. At Coral Cove, 58 transects were sampled before the burial of the Carlin Park reef and 90 after. Eight transects were sampled over natural sand habitats at Carlin Park and six at Coral Cove. An additional 36 hardbottom and 6 sand transects were sampled at Ocean Ridge. Family and species abundances Thirty-six families of fishes were censused among the three hardbottom sites. The most speciose fam- ily was the grunts and margates (Haemulidae) with 11 species of Haemulon and Anisotremus. The wrasses, parrotfishes, and damselfishes (Labridae, Scaridae, and Pomacentridae) had seven, six, and five species, respectively. Four species each of jacks (Carangidae), snappers (Lutjanidae), and clinids (Labrisomidae) were recorded. These seven families contained 50^7^ of the total species censused. Eighty- six taxa (77 identified to species) and 10,491 indi- viduals were censused at all sites (Appendix). At Coral Cove, 64 species and 5093 individuals were recorded. At Carlin Park, 53 species and 4438 indi- viduals were recorded. At Ocean Ridge, 48 species and 960 individuals were recorded. Mean numbers of both species and individuals per transect were similar among all sites (Fig. 2). 512 Fishery Bulletin 97(3). 1999 Table T Mean number of individuals/transect and frequency of occurrence for the most abundant three families, genera, and specie nearshore hardbottom sites. Only predredging data were used for Carlin Park site. CC: Coral Cove (148 transects); CP: Park (112 transects); OR: Ocean Ridge (36 transects); GM: grand mean. sat all Carlin Mean number/transect % frequency Dccurrence CC CP OR GM CC CP OR GM Family Haemulidae 15.5 17.4 9.4 15.5 89 90 92 90 Pomacentridae 5.9 7.9 5.7 6.6 81 95 86 87 Sparidae 5.9 3.7 3.9 37 44 38 Labridae 3.2 3.0 65 64 Genus Haemulon 9.8 15.3 6.2 11.4 75 80 42 75 Stegastes 3.4 6.1 4.3 72 89 73 Anisotremus 5.7 3.2 4.1 74 39 69 Diplodus 5.8 3.7 3.8 35 36 65 Species Haemulon parra 4.4 5.0 3.4 4.5 62 64 33 59 Diplodus argenteus 5.8 3.7 3.8 35 36 36 Stegastes variabilis 5.4 3.7 86 71 Labrisomus nuchipinnis 3.1 2.7 73 69 Abudefduf saxatilis 3.1 2.3 17 31 Anisotremus surinamensis 3.5 2.2 36 43 The three most abundant species were the sailors choice (Haemulon parra), silver porgy [Diplodus argenteus), and cocoa damselfish (Stegastes variablis) with means of 4.5, 3.8, and 3.7 individuals/transect over all sites (Table 1). The most abundant species at Coral Cove, sailors choice, black margate [Anisotremus surinamensis), and hairy blenny [Labrisomus nuchipinnis), represented 32% of all individuals. Seven of the 15 most abundant species at Coral Cove were grunts. At Carlin Park, silver porgy, cocoa damselfish, and sailors choice repre- sented 41% of all individuals. Eight of the 16 most abundant species were grunts. At Ocean Ridge, the most abundant species were silver porgy, sergeant major (Abudefduf saxatilis), and sailors choice. Grunt species ranked first in frequency of occurrence per transect at Coral Cove and Ocean Ridge, and second at Carlin Park (Table 1). Damselfish species ranked first in frequency at Carlin Park and second at the other sites. The most frequently occurring species overall were cocoa damselfish, hairy blenny (Labrisomus nuchipinnus), and sailors choice (Table 1). Normal cluster analysis of samples from all sites resolved three groups that broadly reflected tempo- ral patterns (Fig. 3). No distinct spatial groupings emerged in the normal analysis. Group 1 consisted of 21 samples (eight from Carlin Park, ten from Coral Cove, and three from Ocean Ridge) mostly taken in spring and summer months. Group 2 consisted of 8 samples (four each from Carlin Park and Coral Cove) taken in mid and late summer. Group 3 included the only winter samples (February 1995 and 1996) taken during the project. Inverse cluster analysis revealed seven groups of taxa (Fig. 4). Group A contained 26 common taxa including the most frequently occurring and abun- dant species from visual surveys such as sailors choice, cocoa damselfish, hairy blenny, and silver porgy ( Table 1 ). This group characterized the spring- summer group of samples defined by normal group 1. The remaining six groups consisted of taxa that were temporally variable in their abundance and occurrence in the samples. Group B was character- ized by species that occurred at lower abundances. Groups F and G were represented by single taxa: Apogon maculatus and Archosargus probatocephal us, respectively. The latter species was important in de- fining normal group 3 (Fig. 3). Ordination of samples projected on CA axes 1 and 2 produced a pattern that generally agreed with the normal cluster analysis (Fig. 5A). The eigenvalue for CA axis 1 was 0.218 and accounted for 16. 9*^ of the Lindeman and Snyder: Nearshore hardbottom fishes of southeast Florida 513 variation in the data set, whereas the eigenvalue for CA axis 2 was 0.124 and accounted for 9.77c of the variation in the data set. Samples from August and September at Coral Cove and Carlin Park separated from all other samples along CA axis 1. In general, samples were not spread widely along CA axis 2; however, two samples. May 1995 at Ocean Ridge and February 1995 at Carlin Park, did separate from the other sites. The ordination of taxa on CA axes 1 and 2 showed how the taxa were distributed in relation to the hardbottom samples along these same axes (Fig. 5B). The most common species (e.g. Table 1 ) clustered near the origin of the ordination. Taxa with high scores along CA axis 1 included infrequently occurring spe- cies such as Halichoeres poeyi, Haemulon aurolin- eatum , Mulloidichthys martin Icus. and Caranx ruber. Low scores on CA axis 1 were Echidna catenate, A.canthurus chirurgus, Chactodon occllatus, and Sciaenidae sp. Species with high scores on axis 2 were Sparisoma aurofrenatum, Chaetodon ocellatus, and Sparisoma viride. These were most abundant at Ocean Ridge in May 1995 and were responsible for the sepa- ration of this sample from all others along CA axis 2. In comparisons of hardbottom and natural sand, 20 transects over natural sand plains recorded only four taxa. The clupeid, Harengulajaguana, was most abundant (18 juveniles in two schools total). An uni- dentified Eacinostomus species, Gerres cinereus, and Caranx bartholomaei were also recorded (four, one, and one individuals, respectively). Hardbottom habi- tats typically had over thirty times the individuals per transect as natural sand habitats. Two-sample ^-tests comparing hardbottom with sand habitats rejected the hypothesis of no differences in mean numbers of individuals per transect (P<0.005). Life-stage abundances At all sites, juveniles were the most abundant life stage among the top ten species ( Table 2 ). At the two intensively sampled sites in Jupiter, the numbers of juveniles of all species pooled were significantly greater than any other life stage (ANOVA, P<0.01, Tukey's HSD). There were no significant differences in abundances among newly settled, early juvenile, and adult stages ( Tukey's HSD ). At both Jupiter sites, at least 80% of the individuals were early life stages (pooled newly settled, early juvenile, and juvenile stages )( Fig. 6). Abundances of newly settled and early juvenile stages at Ocean Ridge were similar to the Jupiter sites (Table 2), although numbers of juve- niles and adults were lower. Newly settled stages of over 20 species were recorded on nearshore hard- bottom structures among the three sites. CP_4/94a CP_4/94b CP_5/94 CP_3/95 CP_6/94a CP_4/95b OR_4/95 OR 6/95 OR_5/95 CC_6/95a CC_6/95c CO 4/94 CC_9/96 CC_6/94 CP_6/94b CC_4/95a . CO 6/96 CP_7/94b ■ CC_3/95 004/96 005/96 ■ CP_6/94c CC_6/95b • CC_7/94 ■ CP_7/94a CP_8/94 008/94 CP_9/94 ■ CC_9/94 • CP_2/95 ■ CO 2/96 . Dn Group 1 Group 2 Group 3 0 00 — I — I — r 0 25 -1 — I — I — I — I — I — I 0 50 0 75 Dissimilarity Figure 3 Clustering of visual census samples (sites and times) by UPGMA analysis of a normal Bray-Curtis dis- similarity matrix. CC: Coral Cove: CP: Carlin Park: OR: Ocean Ridge. Eight of the ten most abundant taxa by site were represented primarily by early stages (Table 2). Co- coa damselfish and hairy blenny occurred most abun- dantly as adults. Adults of at least five of the top ten species occurred as residents, not transients. These included sergeant major, hairy blenny, cocoa dam- selfish, silver porgy, and black margate. Adults of a variety of less common species occurred but were of- ten less abundant than early life history stages. Newly settled and juvenile stages often appeared to display more site-fidelity with hardbottom structure than did adults. Six species of grunts (four Haemulon and twoA«/- sotremus) and two species of damselfishes (Stegastes 514 Fishery Bulletin 97(3), 1999 75- 23- 40- 26- 59- 21 - 25- 28- 33- Labnsomus nuchipinnis - 1 Stegasles vanabilis - 4 Halichoeres bivittalus - 3 Anisotremus virginicus - 2 Abudefduf saxatilis - 6 Haemulon parra - 5 Diplodus argenteus - 7 Anisotremus sunnamensis - 8 Haemulon sp (NS) - 11 Haemulon flavolinealum - 13 Haemulon chrysargyreum - 16 Haemulon plumien - 14 Pemphens schomburgki - 20 Haemulon aurolinealum - 24 Acanthunjs bahianus - 9 Equetus acuminalus - 10 Stegasles fuscus - 12 Thalassoma bifasciatum - 19 Lutjanus synagns - 1 7 Haemulon macrostomum - 18 Caranx bartholomaei - 22 Spansoma njbnpinne Acanthunjs chinjrgus Eucinostomus sp Spansoma sp Sciaenid sp (NS) Halichoeres radiatus Halichoeres maculipinna Stegasles sp Kyphosus sp Spansoma aurofrenalum - 6 1 Hypleurochilus sp - 39 Spansoma vinde Scarus vetula Chaetodon ocellatus Odontoscion dentex Lutjanus gnseus Parablennius marmoreus Sphyraena banacuda - 54 Alutenjs scnptus - 57 Coryphoptenjs glaucofraenum - 27 Cantherhines pullus - 29 Malacoctenus tnangulatus - 45 Pomacanthus pani - 32 Diodon hystnx ■ 5 1 Ocyurus chrysurus - 3 1 Scorpaena plumien ■ 42 Caranx ruber - 35 Haemulon sciurus - 44 Haemulon carbonanum - 36 Mulloidichthys martinicus - 37 Halichoeres poeyi - 48 Spansoma chrysopterum Acanthurus coeruleus Psuedupeneus maculatus Echidna catenata Scartella cnstata Gerres cinereus Haemulon melanurum Apogon maculatus Archosargus probalocephalus 47 ■ 50- 56- 30- 58- 49- 55- 53- 52- 0.00 ^ >1 ^^ -T — I — r 0.25 T I I I I I T 0.50 Dissimilarity 0.75 1.00 Figure 4 Clustering offish taxa co-occurrence at three nearshore hardbottom sites by UPGMA analysis of an inverse Bray-Curtis dissimilarity matrix. Numeric codes used in correspondence analysis are next to each name. NS: Newly Settled. Dashed lines delineate groups A-C!. and Abudefduf) ranked within the ten most abun- dant species from all three sites (Table 2). Relative abundances of the life stages of all grunts censused at the Jupiter sites are shown in Figure 7. Early ju- venile stages of the most abundant species, sailors choice, were significantly more abundant than any other life stage at each of the three sites (Kruskal- Wallis ANOVA, P<.001, and a posteriori pairwise comparisons ). Adult sailors choice were significantly lower in abundance than juvenile stages (Kruskal- Wallis ANOVA, P<0.001). Black margate and porkfish, An/.so/rem//.s virginicus, ranked second and Lindeman and Snyder: Nearshore hardbottom fishes of southeast Florida 515 Table 2 Mean number of individuals/transect by life stage for the ten most abundant taxa at each of three s ites. NS: New ly Settled , EJ: Early Juvenile (for haemulids only); J: Juveniles A: Adults. Or ly predredging data were used for Carlin Park site, na = not available. Species Mean number individuals/transect Coral Cove Carlin Park Ocean Ridge 148 transects 112 transects 36 transects NS EJ J A NS EJ J A NS EJ J A Haemulon parra 0.7 2.7 0.8 0.2 0.7 3.8 0.8 0.1 0.4 2.6 0.4 <0.1 Diplodus argenteus 0.3 na 1.7 0.3 1.6 na 4.1 0.1 0.7 na 2.8 0.2 Stegastes variabilis 0.2 na 1.4 1.3 0.4 na 1.4 3.5 0.4 na 0.9 0.6 Halichoeres bivittatus 0.2 na 2.3 0.5 0.3 na 2.1 0.5 0.3 na 1.1 <0.1 Labrisomus nuchipinnis 0 na 0.8 2.3 0 na 0.3 2.0 <0.1 na 0.5 1.8 Abudefduf saxatilis 0.4 na 2.0 0.1 0.3 na 1.1 0.4 1.3 na 1.7 0.1 Anisotremus surinamensis 0.4 1.9 1.1 0.1 0.1 0.3 0.3 0 0.5 0.6 0.4 0 Arjisotremus virginicus 0.9 0.7 0.4 0.2 0.3 0.9 0.3 <0.1 0.9 0.2 0.4 0.3 Haemulon aurolineatum 0 0.7 1.0 0 0 0.1 2.5 0 0 0 0 0 Haemulon sp. 1.8 <0.1 0 0 2.1 <0.1 0 0 0.9 0 0 0 Haemulon flavolineatum <0.1 0.6 0.3 <0.1 0 0.9 1.0 <0.1 0 1.1 0.1 0 Haemulon chrysargyreum 0 <0.1 0.3 0 <0.1 0.3 1.7 0 0 0 0 0 Total means of life stages per site 4.9 6.6 12 5.0 6.3 6.3 15.5 6.6 5 0 5 .} 7 4 3 0 third in overall abundance among grunts and were represented by all life stages (Fig. 7). Tomtate, Hae- mulon aurolineatum , ranked fourth on the basis of large but infrequent influxes of early stages. Outside of these pulses, tomtate was not an abundant or frequently oc- curring species at any site during any life stage. Some newly settled grunts could not be positively identified during visual sui-veys and were pooled as Haemulon sp. (newly settled larvae of Anisotremus are distinctive. Lindeman, 1997a). This group con- tained epibenthic larvae of several species and ranked tenth in abundance among all taxa (Table 2) and fifth among haemulids ( Fig. 7 ). The largest com- ponent of these unidentified schools was probably sailors choice. This assumption is based on 1) the greater relative abundances of sailors choice early juveniles at all sites; 2) the close proximity of sailors choice early juveniles to these newly settled Haemulon sp.; and 3) collections of several newly settled Hae- mulon sp. schools most commonly contained sailors choice upon microscopic examination. Early stages of commercially valuable species oc- curred infrequently during the surveys, although recreationally important species were common. The most abundant commercial family at the nearshore hardbottom sites was the Lutjanidae (snappers). Four snapper species, totaling 58 individuals, were recorded at all sites. Thirty-eight of these were lane snapper, Lutjanus synagris. Thirty-three of these were juveniles, the majority less than five cm. Five newly settled individuals (<2 cm) were also recorded. Ten yellowtail snapper, Ocyurus chrysurus, were re- corded, nine small juveniles and one newly settled individual. Unlike grunts, newly settled and small juvenile snappers were not gregarious, occurring in- dividually or in pairs near interfaces of hardbottom structure and sand. Gray snapper. Lutjanus griseus, and schoolmaster, Lutjanus apodus, occurred only as older juveniles or adults and in low numbers (eight and two individuals, respectively). Comparisons of interannual and seasonal patterns of life stage abundances were limited by temporally unbalanced sampling. Wind and wave conditions from September through February made collection of nearshore visual data in fall and winter erratic or impossible. Several hurricanes during July and Au- gust of 1995 produced wave and turbidity conditions that interrupted summer sampling as well. Other phenomena, including high winds and discharges of turbid water from the Jupiter Inlet, also precluded sampling for extended intervals. Nonetheless, samples were obtained from Coral Cove (the control site) for three consecutive years for the months of April and June. Comparisons of both numbers of spe- 516 Fishery Bulletin 97(3), 1999 cies and numbers of individuals per transect for April among 1994-96 revealed no significant interannual differences (ANOVA,P=0,34; ANOVA,P=0.21). Iden- tical comparisons for June among the same three- year period revealed no differences among mean numbers of individuals (AN OVA, P=0.06), but sig- nificant differences among numbers of species (ANOVA, P<0.05). Pairwise comparisons using Tukey's HSD showed that Coral Cove in June 1995 had significantly more species than in June 1996 (mean species numbers: 8.7 versus 5.3). Seasonal occurrence of only newly settled stages was examined at Coral Cove (Fig. 8). The timing and abundance of species occurrences suggested seasonal variations with peaks of newly settled stages in late spring and early summer. Sizes and numbers of individuals typically increased as summer pro- gressed. Abundances of early stages for most species appeared to be low in February and March prior to peak spawning activity for many species (Garcia- Cagide et al., 1994). However, difficulties in the col- lection of visual sui^veys constrained the examina- tion of fall and winter patterns. 4 -1 A ■ CO 3 - /--■■V5/95 ■"■-.. Group 1 / Q CP ^ OR 2 - '■' \ 6/95 \ 3/95 ^'9^" \ 1 - \ 3/95 e: Group 2 — --.^ ■"JS 4/95\" ^^ 5/96 8/94\ ^ * 4/95b 1 ...---'' 8/94 9/94 B i 0 - 6/95C 5/94 /••■ ,,„,. B ■ 6/96B E ^^^/ 6/95b ^^^^ 4/94B 4/94aE a<"^ ■ U ,/' 1 - 6/95a ■4/96a .•-• \ B6/94c ^'^ S /W"<:'' "■■-- -"" 9/95/ Ep 2/96 \ 2 - \ 6/94b .,•• '■■., \ '\P 2/95) — 'Group 3 Q 1 1 1 1 1 _ 2 -1 0 1 2 3 3 -, 2 - 1 - 0 - -1 - -2 - B 5*42*° 51 43 33 39 ,25 2f' 57 4t 46e ' 56 26 23 U^ 19 13 42^ _ - si 14 5 40 12 S3 43 sz "1 0 Axis 1 Figure 5 (A) Ordination of sample scores on axes 1 and 2 from correspondence analy- sis of a samples by taxa matrix of census data from Coral Cove (CC), Carlin Park (CP), and Ocean Ridge (ORi. Groups 1-3 from Figure 3. (B) Ordination of taxa scores from the same correspondence analysis. Numeric codes for taxa are given in Figure 4. Ichthyofaunal characteristics after habitat burial Prior to habitat burial, fish assemblages at the two Jupiter sites were similar in species composition and relative abun- dance (Figs. 2, 3, and 5A). Pre- and postburial numbers of individuals and species at the control and impact sites are plotted in Figure 9. The hypotheses of no differences in total numbers of in- dividuals and species before and after dredge burial of hardbottom were both rejected (P<0.001) in two sample ^tests for equal variances following a BACIPS design (Stewart-Oaten et al., 1986; Osenberg and Schmitt, 1996). No fishes or exposed hardbottom were recorded in the first postdredging sur- veys at Carlin Park ( 13 April 1995). Sev- eral hardbottom outcroppings ( 1 m high by 3 m wide), parallel to the shore, were exposed at a depth of 1.5 to 2.5 m at the site during the second postdredging sur- veys on 24 April. Two of ten transects crossed narrow outcrops with dense schools of newly settled stages of three grunt species and one drum species (Sciaenidae). Before burial of the reef by extension of the beach width by ap- proximately 60 m, such outcrops were deeper and further offshore than the original hardbottom sampling area. Small outcrops were still present in May and were occupied in two of ten tran- sects by three species of grunts and one damselfish (Fig. 9). Schools of newly settled stages predominated. Such out- crops were not encountered during the 20 surveys in June, and no fishes were present with the exception of several round scad, Decapterus punctatus. Ten Lindeman and Snyder: Nearshore hardbottom fishes of southieast Florida 517 50 m 40 TO TO 30 ■D > C A: All species 20 10 NS EJ J A PE 30 20 TO T3 10 -I B: Grunts ^^£_ NS EJ J A PE Life stage abundances Figure 6 Abundances of different life history stages at the Jupiter hardbottom sites (with 95'7f confidence intervals). A: All species pooled. B: Pooled grunt species only. Only pre- dredging data were used for Carlin Park site. NS: Newly Settled; EJ: Early Juvenile (for grunts only I; J: Juveniles; A: Adults; PE: Pooled Early Stages (=NS+EJ+J). surveys in September 1995 recorded no exposed out- crops or fishes. During the following winter, erosion occurred and the width of the new beach was reduced. Some outcrops were re-exposed by the loss of dredge- fill. However, wind and waves prohibited visual sam- pling during this period. Surveys in February, April, and May of 1996 ( 22 transects total ) recorded no spe- cies (Fig. 9). Discussion Fish assemblages of nearshore hardbottom The diversity of fishes utilizing nearshore hard- bottom habitats of mainland Florida has not been quantified. Qualitative studies by ichthyologists ex- perienced with the substantial taxonomic problems 1000 100 10 1 1000 100 10 1 1000 100 10 1 1000 100 10 1 1000 100 10 1 Newly settled Jjia fEa_ Early juveniles .Lin Juveniles mm Adults Qfl _a_ c?a Total I -fLJsfJ en < a. UJ < I 03 2 < > < 0- 3 iJU m d: D o X I?' < O (3- 5 oj yj OJ LU < $ < < a: < o Figure 7 Comparative abundances of grunts among 12 taxa and 4 life history stages. Data pooled from all Coral Cove sur- veys and predredging Carlin Park surveys (260 transects total). Species represented by abbreviated genus and spe- cies names. within these diverse, largely juvenile assemblages are also lacking. Three studies have included sec- tions on nearshore hardbottom fishes as part of larger project goals. Gilmore (1977) listed 105 species in association with "surf zone reefs" at depths less than two m. Two additional species were added in later papers (Gilmore et al., 1983; Gilmore, 1992). Using visual surveys, Vare (1991) recorded 118 species from nearshore hardbottom sites in Palm Beach County. Futch and Dwinell (1977) included a list of 34 spe- cies obtained from several ichthyocide collections on "nearshore reefs." Including species from these prior studies, 192 species have now been recorded in asso- ciation with nearshore hardbottom habitats of main- land southeast Florida (Table 3.3 in Lindeman, 1997a). Numbers of labrisomid, blenniid, gobiid, and apogonid species may be underestimated owing to their small size or cryptic behaviors. Other hard- bottom habitats of the southeast United States oc- cur in areas with substantially different physi- 518 Fishery Bulletin 97(3), 1999 ographic regimes (Sedberry and Van Dolah, 1984; Chiappone and Sullivan, 1994) and may show dif- fering patterns of fish diversity. Spatial and temporal attributes offish assem- blages at the three sites in the present study were examined by using ordination and cluster analy- sis. Visual census samples collected from March through July were similar in species composition and relative abundance among sites (Fig. 3). This finding is in agreement with the similar plots of individual and species abundances among sites (Fig. 2). The relative homogeneity of these samples was further reflected in the co-occurrence of many taxa including haemulids (Haemulon parra, H. flavolineatum, H. chrysargyreum, Anisotremus virginicus,A. surinamensis), pomacentrids iSteg- astes variabilis, Abudefdufsaxatilis), labrisomids (Labrisomus nuchipinnis), sparids (Diplodus holbrooki), labrids (Halichoeres bivittatus, Thalas- sorna bifasciatum) and scarids (Sparisoma rubri- pinne) (Fig. 4). With the exception of L. nuchi- pinnis and S. variabilis, most taxa occurred as early life stages. Samples from late summer (August and Septem- ber) were distinct from the spring and early sum- mer in both cluster analysis and ordination (group 2, Figs. 3 and 5). The only two samples taken in win- ter (February) differed from all other samples in the analyses (group 3, Figs. 3 and 5). These patterns suggest that some seasonality in assemblage struc- ture existed. This may reflect late spring and sum- mer peaks in larval settlement in contrast to reduced winter settlement and, possibly, influxes of older ju- veniles from inshore lagoonal habitats. Substantial numbers of many species still settled in late sum- mer but were possibly subject to higher predation from older individuals that settled earlier in the year. Various physical disturbances (e.g. winter cold fronts, summer hurricanes) and biological phenomena (variation in larval recruitment) affect the composi- tion of fish assemblages of nearshore hardbottom. The turbidity generated by physical disturbances constrains the visual surveys needed to assess their immediate effects. Nursery habitats and nearshore hardbottom With increasing human modifications of coastal ar- eas, detailed knowledge of habitat usage is a key com- ponent of informed fishery and coastal land manage- ment. Identification of essential habitats includes the evaluation of spatial distributions of structural habi- tats across the shelf and habitat requirements of key taxa. Several lines of evidence suggest that nearshore hardbottom habitats along the mainland coast of east 25 n ~ 20 o 15 10 Newly settled stages I I I I I T-T I I I I AMJJASONDJFMAMJJASONDJFMAMJ 1994 1995 1996 Figure 8 Mean number of newly settled individuals per transect for all species pooled at Coral Cove. 1994-1996 In =148 transects). Florida can serve as nursery areas for many coastal fish species. Over 80'v^ of the individuals at all sites were early life stages. Eight of the top ten species were consistently represented by early stages. Use of hardbottom habitats was recorded for newly settled stages of more than 20 species. In addition, other natural habitats with substantial vertical relief were absent from the shallow physiographic regimes where nearshore hardbottom occurred. Although suggestive of nursery value, these lines of evidence need to be viewed in the appropriate con- text. High abundances of early life stages compared with adults do not guarantee that a habitat is a valu- able nursery. High mortality rates in many reef fish populations (Sale, 1980; Shulman and Ogden, 1987; Richards and Lindeman, 1987; Jones, 1991) suggest that early stages will typically be more abundant than adults. If spatial distributions of all life stages are homogeneous, all habitats will have more early stages than adults. However, the abundances of early stages on nearshore reefs probably reflect more than just larger numbers of homogeneously distributed recruits. Newly settled stages of eight of twelve spe- cies of grunts and eight of nine species of snappers of the southeast mainland Florida shelf have been recorded primarily in depths less than ten m (Linde- man et al., 1998). Adults of most species are typi- cally uncommon or absent in shallow habitats. There is considerable evidence for cross-shelf habitat seg- Lindeman and Snyder Nearshore hardbottom fishes of soutfieast Florida 519 (0 regation among life stages of many grunt and snapper species from other regions as well; early demersal stages appear to most commonly use shallow habitats (Starck, 1970; Dennis, 1992). Similar ontogenetic differences in distribution and abundance exist for many other taxa that utilize nearshore hardbottom habitats. Determining if the availability of habitat structure limits survival of early stages is important in assessing nursery value. Ab- sences of habitat structure can result in increased predation or lowered growth (Hixon, 1991). In southeast mainland Florida, many natural nearshore marine habitats outside of coastal lagoons and be- tween 25°30'N and 26°20'N (Dade and Broward Counties) are sand plains lacking hardbottom and substantial three-dimen- sional structure (ACOE, 1996). Although large stretches of nearshore hardbottom exist between 26°20'N and 27°50'N (Palm Beach, Martin, St. Lucie, and Indian River Counties) these habitats are often separated by kilometers of sand plains. There are no other natural habitats in the same near- shore areas that can support equivalent abundances of early life stages. These con- ditions could promote a demographic bottle- neck that limits local adult populations owing to limited habitat availability for early stages. Despite their shallow depth, nearshore hardbottom reefs are positioned within cur- rent and tide regimes that can support con- siderable larval abundances. The occur- rence of presettlement larvae in these ar- eas is reflected by the abundances of newly settled stages in the present study and larvae in nearshore zones of Gulf of Mexico barrier islands (Ruple, 1984; Ross et al., 1987). Newly settled individuals were not recorded during any surveys of pure sand habitats in the present study. However, the presence of nearshore hardbottom promoted substantial coloni- zation of shallow outcrops by larvae of many spe- cies, including haemulids, lutjanids, sparids, labrids, gerreids, sciaenids, and scarids. Ecotones with high vertical relief (e.g. hardbottom-sand interfaces near ledges) sometimes had large aggregations of newly settled stages of these taxa. However, microhabitat- scale distributions of fishes on nearshore hardbottom remain unquantified. Use of nearshore hardbottom reefs as nurseries may be bidirectional across the shelf Both inshore and offshore migrations during differing ontogenetic I I Coral Cove (control) u |TT A i ! Carlin Park (impact) AMJ JASONDJ FMAMJ JASONDJ I I I I I FMAMJ 1994 1995 1996 20 -r 15 - 10 5 - 10 - 15 20 I 11 f! I h Coral Cove il |i I ! I A SSI ts Carlin Park AMJ JASONDJ FMAMJ "T \ ] 1 I ! [ : ' \ ! r" JASONDJ FMAMJ Figure 9 Mean numbers of individuals and species at control and impact sites m Jupiter, FL. Arrows indicate timing of dredge burial of hardbottom reef stages can be facilitated by habitats positioned cen- trally on the shelf Nearshore hardbottom may serve a primary nursery role for incoming early life stages that would undergo increased predation mortality without shelter. Nearshore hardbottom may also serve as secondary nursery habitat for juveniles that emigrate out of inlets towards offshore reefs. This pattern is seen in gray snapper and bluestriped grunt which often settle inside inlets and primarily use nearshore hardbottom as older juveniles. In addition, some species use these structures as resident nurs- eries, settling, growing-out, and maturing sexually as permanent residents (e. g. pomacentrids, labri- somids). A secondary nursery role may also result from increased growth because of higher food avail- abilities in structure-rich environments. The inter- mediate cross-shelf positioning and other attributes reviewed above suggest nearshore hardbottom rep- 520 Fishery Bulletin 97(3), 1999 resents essential fish habitat for many species fol- lowing NOAA (1996). Bidirectional use of nursery habitats positioned between inshore grassbeds and offshore reefs requires further study. From abundance patterns of early life stages and the absence of any nearby natural habitats with high vertical relief, nearshore hardbottom of southeast mainland Florida was estimated to have nursery value for 34 species (Appendix). Empirical correla- tion of variation in early survival with adult popula- tion size is an important but rarely achieved compo- nent of nursery area evaluation. Combining experi- mental studies of habitat requirements with broad field surveys can aid in connecting organism-scale attributes with population-scale patterns (Serafy et al., 1997). Early demersal stages of several of the most representative taxa of nearshore hardbottom (e.g. grunt and damselfish species) can be collected and manipulated in the field and laboratory with relative ease (Lindeman, 1986; 1997a). These taxa may serve as useftil models for nursery habitat studies that ex- perimentally assess habitat requirements. Effects of dredge-and-fill activities on ichthyofauna Burial of the nearshore hardbottom habitat at Carlin Park with dredged sand significantly lowered the abundances of both species and individuals (Fig. 9). Before burial, 54 species were recorded, with mean abundances of 38 individuals and 7.2 species per transect (?! = 112 transects). After burial, eight spe- cies were recorded with mean abundances of less than one individual and species per transect in=92 transects). No quantitative studies on the effects of nearshore hardbottom burial on fishes are available in the peer- reviewed literature for comparison. The final supplemental environmental impact statement (EIS) for the Carlin project (Palm Beach Co. Dep. Environ. Resources Management, 1994) summarized several agency and contractor surveys between 1985 and 1990 at Carlin Park. Ten to forty- eight fish species were recorded from qualitative surveys of the hardbottom. Statements regarding the habitat value of nearshore reefs and dredging effects in the Carlin Park EIS emphasized the variable na- ture of reef exposure and forecast that fish impacts would be minimal and temporary. Primary impacts predicted for fishes were 1) short-term displacement during construction; and 2) temporary loss of food sources. The EIS also emphasized that impacts would be reduced by several features of the project design and nearshore environment. These features included the following: 1) the fishery value of impacted spe- cies was low; 2) some amount of hardbottom would remain or would be constructed for mitigation if needed; and 3) construction of the project would take place when fish populations were at their lowest. No mention of direct or indirect mortality upon fishes was made. The biological assumptions within this EIS are similar to those found in related documents (e.g. ACOE, 1996). For the following reasons, it is sug- gested that some of these assumptions may be tenu- ous. The majority of individuals displaced by hardbottom burial in southeast Florida are early stages of economically and ecologically valuable spe- cies (Appendix; Figure 9). Early demersal life stages are particularly vulnerable to predators (e.g. Shulman and Ogden, 1987). Displacement was per- manent for most individuals because almost all prior habitat was eliminated for at least 15 months (the postburial duration of the present study). Because of behavioral and morphological constraints on flight responses, high mortalities are probably unavoidable for many cryptic species, newly settled life stages, or other site-associated taxa subjected to direct habi- tat burial (Table 4.10 in Lindeman, 1997a). Whether a fish population is seasonally low at the time a project begins is insignificant if dredging will bury the habitat immediately before the peak period of larval settlement,^ as in the Carlin Park project. In addition, loss of reef-associated food sources was probably substantial over this period. No substantial habitat structure was present within at least 0.8 km of the Carlin Park reef during its burial. The closest natural structure was east- ward at depths of at least 10 m. These deeper midshelf habitats may be utilized by relatively few grunt and snapper species during the newly settled and early juvenile stages. To the south, no substan- tial hardbottom was present for at least 4 km. To the north, the jetties of the Jupiter Inlet were approxi- mately 2 km away. However, fishes in a northerly flight response had to negotiate a water column with zero visibility because dredge fill was dumped north- to-south. Any early stages of fish reaching the jet- ties would probably encounter high predation from older piscivores utilizing the large cavities among the armor-stone boulders of the artificially deepened jetty area (Lindeman, 1997a). A postburial mitigation project using shallow arti- ficial reefs of limestone boulders was proposed in the ^ Hackney, C. T., M.H.Posey, and S.W. Ross. 1996. Summary and recommendations. In C. T. Hackney, M. H. Posey, S. W. Ross and A. R. Norris (eds.l, A review and synthesis of data on surf zone fishes and invertebrates in the south Atlantic Bight and the potential impacts from beach renourishment, p. 108- 111. Rep. to U. S. Army Corps of Engineers. Wilmington Dis- trict, Wilmington. NC. Lindeman and Snyder: Nearshore hardbottom fishes of soutfieast Florida 521 Carlin Park EIS. In the summer of 1998, three years after the burial, construction of approximately 1.6 ha of mitigation reefs began. If constructed before burial and at similar depths, mitigation reefs may have provided a refuge for a sizeable fraction of the thousands of displaced fishes during the burial of the hardbottom reef, as well as thousands of subsequent new recruits. Even with prompt construction of arti- ficial reefs, many factors can limit the net produc- tion of biomass (Grossman et al., 1997). Some bur- ied outcroppings were uncovered because of erosion of the project fill. However, structural support for two years of larval recruitment, shelter from post- settlement predation, and food for growth, were prob- ably eliminated at the hardbottom burial site. Nearshore hardbottom areas, such as Carlin Park, can be exposed to extended periods of wave energy and turbidity, particularly during winter months. However, conditions in winter do not dilute the po- tential significance of artificial burial during the spring and summer months. These are the periods of peak usage of hardbottom habitats by newly settled and juvenile stages of fishes. In the absence of dredg- ing, nearshore areas typically show high reef expo- sures and reductions in physiological stressors dur- ing the spring-summer recruitment window. Elimi- nation of this recruitment window by habitat burial for one or more years, regardless of winter dynam- ics, may substantially degrade the value of the pri- mary natural nursery habitats along the windward shorelines of Florida's east coast. The above reasons suggest a risk-averse approach to hardbottom burial, as previously suggested for invertebrate fauna (Nelson, 1989). The cumulative effects on fishes of repeated burial of nearshore habitats and other byproducts of these projects remain unknown. Cascading disturbances with ecosystem-scale effects can be hypothesized for a number of cumulative anthropogenic modifications in south Florida (e.g. Butler et al., 1995; Ault et al., 1998). Habitats affected by dredging or filling can show effects over temporal and spatial scales that are rarely considered (Vestal and Rieser. 1995; Lindeman, 1997b). For example, chronically elevated turbidities could lead to declines in primary produc- tion for frequently dredged areas of the southeast Florida shelf. Conclusive statements on the cumula- tive effects of large-scale dredging upon fishes will ultimately depend on the correlation of variations in early survival with adult population sizes, a rarely achieved task, even when effects may be substantial (Osenberg and Schmitt, 1996). However, the current absence of basic information on both short- and long- term scales can also be treated as an opportunity. Large dredge projects affecting midshelf and near- shore habitats will continue along the southeast Florida shelf at one- or two-year intervals. Basic questions on dredge-and-fill effects upon habitat use, predation, and growth, await study within a diverse assemblage of nearshore fishes. Acknowledgments J. Ault, J. Bohnsack, G. Dennis, G. Gilmore, P. Glynn, M. Harwell, and H. Wanless provided substantial review comments. Several anonymous reviewers were also very helpful. Conversations with the late David Kirtley on sabellariid reefs were consistently valuable. The assistance of these agency personnel is acknowledged: P. Davis, D. Ferrill, J. Iliff A. Mager, and C. Sultzman. Assistance was also provided by J. Gonzalez, B. Hartig, R. Hudson, C. Leyendecker, M. Perry, R. Pugliese, M. Ridler, P. Sale, E. Schoppaul, J. Serafy, A. Stone, K. Snyder, G. Waugh, and D. Wilder. Funding was provided by the Elizabeth Ordway Dunn Foundation, the American Littoral Society, Coastal Research and Education, Inc., the South Atlantic Fishery Management Council, and the Dorr Foundation. Literature cited ACOE (Army Corps of Engineers). 1996. Coast of Florida erosion and storm effects study: re- gion III with final environmental impact statement. ACOE Tech. Rep. Jacksonville District. Three volumes and appendices A-I. Ault, J. S., J. A. Bohnsack, and G. Meester. 1998. A retrospective 1 1979-19961 multispecies assessment of coral reef fish stocks in the Florida Keys. Fish. Bull. 96(31:395-414. Bray, J. R., and J. T. Curtis. 1957. An ordination of the upland forest communities of southern Wisconsin. Ecol. IMonogr 27:32.5-349. Briggs, J. C. 1974. Marine zoogeography. McGraw-Hill. New York, >A', 475 p. Butler rV, M. J., J. H. Hunt, W. F. Herrnkind, M. J. Childress, R. Bertelsen, W. Sharp, T. Matthews, J. M. Field, and H. G. Marshall. 1995. Cascading disturbances in Florida Bay. U.S.A.: cyanobactena blooms, sponge mortality, and implications for juvenile spiny lobsters Panulirus argus. Mar. Ecol. Prog. Ser 129:119-125. Chiappone, M., and K. M. Sullivan. 1994. Ecological structure and dynamics of nearshore hard- bottom communities in the Florida Keys. Bull. Mar. Sci. .54(31:747-756. CuUiton, T. J., M. A. Warren, T. R. Goodspeed, D. G. Remer, C. M. Blackwell, and J. J. McDonough. 1990. Fifty years of population change along the nation's coasts, 1960-2010. Second Rep. Coastal Trends Series. Strat. Assess. Branch. NOAA. 41 p. 522 Fishery Bulletin 97(3), 1999 Dennis, G. D. 1992. Resource utilization by members of a guild of benthic feeding coral reef fish. Ph.D. diss., Univ. of Puerto Rico, Mayaguez, Puerto Rico, 224 p. Duane, D. B., and E. P. Meisburger. 1969. Geomorphology and sediments of the nearshore con- tinental shelf, Miami to Palm Beach, Florida. USACOE Coastal Engineering Center, Tech. Memo. No. 29, 47 p. Futch, C. R., and S. E. Dwinell. 1977. Nearshore marine ecology at Hutchinson Island, Florida: 1971-1974. IV. Lancelets and fishes. Fla. Mar J Res. Publ. No. 24, 23 p. Garcia-Cagide, A., R. Claro, and B. V. Koshelev. 1994. Reproduccion. /;; R. Claro (ed.), Ecologiade los peces niarinos de Cuba, p. 187-262. Centro de Investigaciones de Quintana Roo, Mexico, 525 p. Gauch, H. G. 1982. Multivariate analysis in community ecology. Cam- bridge Univ. Press, Cambridge, 298 p. Gilmore, R. G., Jr. 1977. Fishes of the Indian River Lagoon and adjacent wa- ters, Florida. Bull. Fl. St. Mus. Bio. Sci. 22(3), 147 p. 1992. Striped croaker, Bairdiella sanctaeluciae. In C. R. Gilbert (ed.). Rare and endangered biota of Florida. II: Fishes, p. 218-222. University Press of Florida, Gainesville, FL, 242 p. 1995. Environmental and biogeographic factors influenc- ing ichthyofaunal diversity: Indian River Lagoon. Bull. Mar. Sci. 57(11:153-170. Gilmore, R. G., P. A. Hastings and D. J. Herrema. 1983. Ichthyofaunal additions to the Indian River lagoon and adjacent waters, east-central Florida. Fla. Sci. 46:22-30. Goldberg, W. M. 1973. The ecology of the coral-octocoral communities off the southeast Florida coast: geomorphology, species composi- tion, and zonation. Bull. Mar Sci. 23(31:465-488. Grossman, G. D., G. P. Jones, and W. Seaman Jr. 1997. Do artificial reefs increase regional fish production? A review of existing data. Fisheries 22(4):17-23. Hixon, M. H. 1991. Predation as a process structuring coral reef fish communities. In P. F. Sale (ed.),The ecology of fishes on coral reefs, p. 475-500. Academic Press, San Diego, CA, 754 p. Hoffmeister, J. E. 1974. Land from the sea: the geologic story of south Florida. Univ. Miami Press. Coral Gables, FL, 143 p. Humann, P. 1994. Reef fish identification: Florida, Caribbean, Bahamas. New World Press, Jacksonville, FL, 396 p. Jongman, R. H. G., C. J. F. ter Braak, and O. F. R. van Tongeren (eds). 1995. Data analysis in community and landscape ecology. Cambridge Univ. Press, Cambridge, 299 p. Jones, G. P. 1991. Postrecruitment processes in the ecology of coral reef fish populations: a multifactorial perspective. In P. F. Sale (ed. I, The ecology of fishes on coral reefs, p. 294-330. Aca- demic Press, San Diego, CA, 754 p. Kirtley, D. W. 1994. A review and taxonomic revision of the family Sabel- lariidac, Johnston, 1865 (Annelida; Polychaeta). Sabecon Press Science Series 1. Vero Bch., FL, 223 p. Kirtley, D. W. and W. F. Tanner. 1968. Sabellariid worms: builders of a major reef type. J. Sed. Petrol. 38(l):73-78. Lindeman, K. C. 1986. Development of larvae of the French grunt. Haemulon flavolineatum. and comparative development of twelve western Atlantic species of Haemulon (Percoidei, Haemu- lidae). Bull. Mar Sci. 39(31:673-716. 1997a. Development of grunts and snappers of southeast Florida: cross-shelf distributions and effects of beach man- agement alternatives. Ph.D. diss., Univ. Miami, Coral Gables, FL, 419 p. 1997b. Comparative management of beach systems of Florida and the Antilles: applications using ecological as- sessment and decision support procedures. In G. Cam- bers (ed.). Managing beach resources in the smaller Car- ibbean islands, p. 134-164. UNESCO Coastal Region and Small Island Paper 1, 269 p. Lindeman, K. C, G. A. Diaz, J. E. Serafy, and J. S. Ault. 1998. A spatial framework for assessing cross-shelf habi- tat use among newly-settled grunts and snappers, Proc. Gulf Carib. Fish. Inst. 50:385-416. NRC (National Research Council). 1995. Beach nourishment and protection. National Acad- emy Press, Washington, DC, 334 p. Nelson, W. G. 1989. Beach nourishment and hard bottom habitats: the case for caution. In S. Tait (ed.), Proc. 1989 National Conf Beach Preserv. Technol., p. 109-116. Fl. Shore and Beach Preserv. Assoc, Tallahassee, FL. Nelson, W. G., and L. Demetriades. 1992. Peracariids associated with sabellariid worm rock iPhragmatopoma lapidosa Kinberg) at Sebastian Inlet, Florida, U.S.A. J. Crust. Biol. 12(4):647-654. NOAA (National Oceanic and Atmospheric Administration). 1996. Magnusen-Stevens Fishery Conservation and Man- agement Act, as amended through Oct. 1 1 , 1996. U.S. Dep. Commer, NOAA Tech. Memo. NMFS-F/SPO-23, 121 p. Osenberg, C. W., and R. J. Schmitt. 1996. Detecting ecological impacts caused by human activities. In R. J. Schmitt and C. W. Osenberg (eds.), Detecting ecological impacts, p. 3-15. Academic Press, San Diego, CA, 401 p. Palm Beach County Dept. Environ. Resources Management. 1994. Palm Beach County, Florida, shore protection project, from Martin County line to Lake Worth Inlet and from south Lake Worth Inlet to Broward County line: Jupiter/ Carlin segment. Final supplemental environmental im- pact statement. Submitted to L'SACOE, Jacksonville Dis- trict Office, 80 p. with appendices. Richards, W. J., and K. C. Lindeman. 1987. Recruitment dynamics of reef fishes; planktonic pro- cesses, settlement and demersal ecologies, and fishery analysis. Bull. Mar Sci. 41(2):392-410. Robins, C. R., and G. C. Ray. 1986. Afield guide to Atlantic coast fishes of North America. Houghton Mifflin Co., Boston, MA, 354 p. Rohlf, F. J. 1997. NTSYS-pc: numerical taxonomy and multivariate analysis .system, version 2.0. Exeter Publishing, Setauket, NY, iil p." Ross, S. T., R. H. McMichael Jr., and D. L. Ruple. 1987. Seasonal and diel variation in the standing crop of fishes and macroinvertebrates from a Gulf of Mexico surf zone. Estuarine Coastal Shelf Sci. 25:391-412. Ruple, D. L. 1984. Occurrence of larval fishes in thesurf zoneof a north- em Gulf of Mexico barrier island. Estuarine Coastal Shelf Sci. 18:191-208. Lindeman and Snyder: Nearshore hardbottom fishes of southeast Florida 523 Sale, P. F. 1980. The ecology of fishes on coral reefs. Oceanogr. Mar. Biol. 18:367-421 Sedberry, G. R., and R. F. Van Dolah. 1984. Demersal fish assemblages associated with hard-bot- tom habitat in the South Atlantic Bight of the U. S, A. Environ. Biol. Fishes ll(4):'241-258. Serafy, J. E., K. C. Lindeman, T. E. Hopkins, and J. S. Ault. 1997. Effects of freshwater canal discharges on subtropi- cal marine fish assemblages: field and laboratory observations. Mar. Ecol. Prog. Sen 160:161-172. Shulman, M. J., and J. C. Ogden. 1987. What controls tropical reef fish populations: recruit- ment or benthic mortality? An example in the Caribbean reef fish, Haemulon flavolineatum. Mar. Ecol. Prog. Ser. 39:233-242. Sneath, P. H. A., and R. R. Sokal. 1973. Numerical taxonomy, the principles and practice of numerical classification. W.H. Freeman and Co., San Francisco, CA, 573 p. Starck, W. A. 1970. Biology of the gray snapper, Lutjanus griseus (Lin- naeus), in the Florida Keys. Stud. Trop. Oceanogr. Univ. Miami 10:11-150. Stewart-Oaten, A. 1996. Goals in environmental monitoring. In R. J. Schmitt and C. W. Osenberg (eds.l. Detecting ecological impacts, p. 17-26. Academic Press. San Diego, CA, 401 p. Stewart-Oaten, A., W. W. Murdoch, and K. R. Parker. 1986. Environmental impact assessment: "pseudore- plication" in time? Ecology 67:929- 940. ter Braak, C. J. F. 1988. CANOCO— a FORTRAN program for canonical com- munity ordination. Microcomputer Power, Ithaca, NY, 95 p. Vare, C. N. 1991. A survey, analysis, and evaluation of the nearshore reefs situated off Palm Beach County, Florida. M.S. the- sis. Florida Atlantic L'niv., Boca Raton, FL, 165 p. Vestal, B. and A. Rieser. 1995. Part I - Syntheses, with annotated bibliography. In: Methodologies and mechanisms for management of cumu- lative coastal environmental impacts. NOAA Coastal Ocean Program Decision Analysis Series No. 6. NOAA Coastal Ocean Office, Silver Springs, MD, 139 p. 524 Fishery Bulletin 97(3), 1999 Appendix Total abundances of all species visually surveyed at three nearshore hardbottom sites, southeast Florida Only predredgin I data were used for Carlin Park site. * = hypothesized to use nearshore hardbottom as a nursery habitat Isee discussion). Coral Carlin Ocean Rank Common name Species Cove Park Ridge Total 1 Sailors choice Haemulon parra* 649 555 122 1326 2 Silver porgy Diplodus argenteus* 344 647 132 1123 3 Cocoa damselfish Stegastes variabilis* 420 600 66 1086 4 Slippery dick Halichoeres bivittatus* 439 327 50 816 5 Hairy blenny Labnsomiis nuchipmnis* 463 262 81 806 6 Sergeant major Abudefduf saxatilis* 367 199 112 678 7 Black margate Anisotremus surinamensis* 513 68 55 636 8 Porkfish Anisotremus virginicus* 331 174 61 566 9 Tomtate Haemulon aurolineatum* 245 295 8 548 10 Grunt sp. Haemulon sp.* 266 233 34 533 11 French grunt Haemulon flavolineatum* 134 210 43 387 12 Smallmouth grunt Haemulon chrysargyreum* 60 222 10 292 13 White grunt Haemulon plumieri* 70 150 1 221 14 Glassy sweeper Pempheris schomburgki* 153 21 32 206 15 Dusky damselfish Stegastes fuscus* 75 83 9 167 16 High-hat Equetus acuminatus* 54 59 13 126 17 Ocean surgeon Acanthurus bahianus* 51 12 17 80 18 Doctorfish Acanthurus chirurgus* 63 2 7 72 19 Redfin parrotfish Spansoma rubripinne* 52 14 2 68 20 Mojarra sp. Eucmostomus sp. 37 20 2 59 21 Spanish grunt Haemulon macrostomum* 14 35 1 . 50 21 Yellow jack Caranx bartholomaei 9 41 0 50 23 Yellow goatfish Mulloidichthys martimcus 34 8 0 42 24 Lane snapper Lutjanus synagns* 23 12 3 38 25 Bluehead wrasse Thalassoma bifasciatum 22 7 7 36 25 Croaker sp. Sciaenid sp. 22 14 0 36 27 Redtail parrotfish Sparisoma chrysopterum * 16 14 3 33 28 Damselfish sp. Stegastes sp. 9 5 18 32 29 Parrotfish sp. Spansoma sp. 14 14 0 28 30 Reef croaker Odontoscion dentex* 13 3 8 24 30 Bar jack Caranx ruber 2 20 2 24 32 Chubsp. Kyphosus sp. 10 4 9 23 33 Bridled goby Coryphopterus gla ucofraen u m 2 19 1 22 34 Clown wrasse Halichoeres maculipmna * 8 5 4 17 35 Anchovy sp. Engraulid sp. 15 0 0 15 36 Puddingwife Haliehoeres radtatus* 4 6 4 14 36 Orangespotted filefish Cantherhines pull us 2 11 1 14 38 French angelfish Pomacanthus paru* 5 5 3 13 39 Seaweed blenny Parablennius marmoreus 2 5 5 12 40 Caesar grunt Haemulon carbonarium* 3 7 1 11 41 Yellowtail snapper Ocyurus chrysurus* 3 5 2 10 41 Striped croaker Bairdiella sanctcluciae* 10 0 0 10 43 Stoplight parrotfish Sparisoma viride 4 1 4 citn 9 linurcl Lindeman and Snyder: Nearshore hardbottom fishes of southieast Florida 525 Appendix (continued) Coral Carlin Ocean Rank Common name Species Cove Park Ridge Total 43 Redband parrotfish Sparisoma aurofrenatiim 1 0 8 9 44 Gray snapper Lutjanus griseus 6 0 2 8 44 Porgy sp. Sparid sp. 0 8 0 8 44 Bluestriped grunt Haemulon sciurus 2 4 2 8 44 Spanish sardine Sardirwila aiinta 8 0 0 8 49 Molly miller Scartella cnstata* 2 5 0 7 49 Blackear wrasse Halichoeres poeyi 6 1 0 7 51 Sheepshead Archosargus probatocephalus 1 5 0 6 51 Blue tang Acanthurus coeruleus* 3 3 0 6 51 Spotted goatfish Pseudupeneus maculatus 3 0 3 6 54 Saddled blenny Malacoctenus tnangulatus* 1 3 1 5 55 Barbfish Scorpaena plumieri 2 2 0 4 55 Queen parrotfish Scarus vetula 0 1 3 4 57 Flamefish Apogon maculatus 0 3 0 3 57 Yellowfin mojarra Gerres cinereus 3 0 0 3 57 Blue runner Caranx crysos 3 0 0 3 57 Spotfin butterflyfish Chaetodon ocellatus 0 1 2 3 61 Balloonfish Diodnn hyslrix 1 0 1 2 61 Chain moray Echidna catenata 2 0 0 2 61 Scrawled cowfish Lactophrys quadncornis 1 0 1 2 61 Schoolmaster Lutjanus apodus 2 0 0 2 61 Blenny sp. Bleniid sp. 2 0 0 2 61 Cottonwick Haemulon melanurum 2 0 0 2 61 Great barracuda Sphyraena barracuda 1 0 1 2 61 Scrawled filefish Aluterus scriptus 1 0 1 2 69 Bicolor damselfish Stegastes partitus 0 0 1 69 Orangespotted goby Nes longus 0 1 0 69 Spanish hogfish Bodianus rufus 0 1 0 69 Spotted snake eel Myrichthys acuminatus 0 1 0 69 Gray angelfish Pomacanthus arcuatus* 1 0 0 69 Sharpnose puffer Canthigaster rostrata 1 0 0 69 Greater soapfish Rypticus saponaceus 1 0 0 69 Smooth trunkfish Lactophrys triqueter 1 0 0 69 Hogfish Lachnolaimus maxim us 0 1 0 69 Puffcheck blenny Labrisomus bucciferus 1 0 0 69 Nurse shark Ginglymostoma cirratum 0 1 0 69 Squirrelfish Holocentrus rufus 0 1 0 69 Blue angelfish Holacanthus bermudensis* 0 0 1 69 Rosy blenny Malacoctenus macropus 0 1 0 69 Spotted moray Gymnothorax moringa 1 0 0 69 Goldentail moray Muraena miliaris 0 1 0 69 Atlantic spadefish Chaetodipterus faber 1 0 0 69 Sand drum Umbrina coroides 1 0 0 Total taxa 72 60 50 86 Total individuals 5093 4438 960 10491 526 Abstract.— Growth and sexual matu- ration were studied in the ghost shrimp Lepidophthalmus sinuensts, a pest spe- cies infesting oligohaline penaeid shrimp culture ponds on the Caribbean coast of Colombia. Sex ratio was signifi- cantly female-biased over four years of sampling. Development of ovaries, in- dexed as relative width, peaked prior to peak occurrence of ovigerous females and either coincided with or immedi- ■' ately followed the quarter of lowest ambient salinity. Ovigerous females oc- curred in all months, but the greatest mean percentage occurred in the first or second quarter of each year. Size and coloration of ovaries varied by matura- tional stage, and monthly counts of eggs per female peaked in February-April. Detection of recruits <8-10 mm cara- pace length (CD was sporadic, and tracking of growth in recruitment co- horts was not possible. Mean CL of the population increased slowly over the first two years of sampling, as the per- centage of males in the population in- creased. Sexual maturity in both males and females was evident in relative growth changes of the major chela. Analysis of growth in chela width, scaled to carapace length, suggested that males mature at about 11.0-11.3 mm CL, whereas females mature near 10.8-11.2 mm CL. Prematuration posi- tive allometric growth did not differ be- tween sexes, whereas postmaturation negative allometric growth in females differed significantly from strongly positive allometric growth in postma- turation males. Relative growth in wet weight of postmaturation males signifi- cantly exceeded that of postmaturation females. Expected female weight accu- mulations from episodic ovarian devel- opment were perhaps offset by simulta- neous negative postmaturation growth of the female major chela. An understand- ing of growth, maturation, and the re- productive cycle of this shrimp species will help develop management strate- gies to control infestations by this callianassid pest species. Growth and maturation of the ghost shrimp Lepidophthalmus sinuensis Lemaitre and Rodrigues, 1991 (Crustacea, Decapoda, Callianassidae), a burrowing pest in penaeid shrimp culture ponds* Sergio F. Nates Darryl L. Felder Department of Biology and Laboratory lor Crustacean Research, University of Soutfiwestern Louisiana Lafayette, Louisiana 70504-2451 E-mail address (for D L Felder, contact author) dlf4517'fflusl edu Manuscript accepted 11 August 1998. Fish, Bull. 97:526-541 11999). In northern South America, the ghost shrimp Lepidophthalmus sinuensis Lemaitre and Rodrigues has been found restricted to oUgo- haUne estuaries on the Caribbean coast of Colombia (Lemaitre and Rodrigues, 1991). From previous study of a congeneric species, osmo- regulatory capacity and toler- ance of hypoxia in this genus (Felder, 1978, 1979; Felder et al., 1986) appear to exceed abilities re- ported in most other callianassid genera (Thompson and Pritchard, 1969; Forbes, 1974, 1978; Mukai and Koike, 1984; Posey, 1987 ), Such adaptation has allowed members of the genus to exploit extensively low salinity habitats, especially richly organic silty bars and banks of lower river mouths or intertidal mudflats (Felder and Manning, 1997), In L. sinuensis and some eastern Pacific congeners (Nates and Felder, 1998), this adaptation also includes sediments of tropical, estuarine ponds in which penaeid shrimp are cultured. Dense accumulations of ghost shrimp in these penaeid culture ponds are favored by their abbrevi- ated larval development (Nates et al., 1997). At high levels of infesta- tion by these burrowers, metabolic impacts and bioturbation appear to decrease penaeid shrimp production by reducing survival rates of post- larvae and slowing growth to mar- ketable size (Nates and Felder, 1998 ). A thorough understanding of maturation and reproductive cycles in L. sinuensis is essential to devis- ing effective long-term control mea- sures for such infestations, espe- cially to limit or replace the pesti- cide treatments that have been used to date (Nates and Felder, 1998), Thus far, timing and frequency of pesticide applications in shrimp mariculture have been devised by trial and error effects on penaeid pro- duction, without specific attempts to target treatments to peaks in repro- ductive activity of ghost shrimp or to achieve control by modifications in water management. Although some species of the Callianassidae burrow to depths of 2 m or more (Pohl, 1946; Poore and Suchanek, 1988; Manning and Felder, 1991; Felder and Griffis, 1994), the difficulty of monitoring many natural populations has been overcome by collecting them with yabby pumps (see Manning, 1975). Several aspects of population biol- ogy for varied genera are now well documented (Pohl, 1946; Hailstone and Stephenson, 1961; Phillips, * Contribution 57 from the University of Southwestern Louisiana, Laboratory of Crustacean Research, Lafayette, LA 70.504-2451. Nates and Felder: Growth and maturation of Lepidophthalmus sinuensis 527 1971; Poore and Suchanek, 1988; Witbaard and Duineveld. 1989; Tamaki and Ingole, 1993; Rowden and Jones, 1994, 1995; Dumbauld et al., 1996), and the importance of callianassids in sediment turnover and nutrient cycling has been noted (Rowden and Jones, 1993; Ziebis et al., 1996). Ecological impacts oi Lepidophthalmus spp. can dominate those of other estuarine infauna, as noted in recent studies of the warm-temperate species L. louisianensis in the northern Gulf of Mexico (Felder and Lovett, 1989; Felder and Griffis, 1994) and the tropical species L. sinuensis on the Caribbean coast of Colombia (Nates and Felder, 1998). Larval life histories for both L. louisianensis and L. sinuensis have been described (Nates et al., 1997), but later life stages and matura- tion remain little known for the tropical species L. sinuensis. Objectives of the present study were to analyze the periodicity and frequency of reproduc- tive activity, to determine rates of growth in mor- phological features, and to define allometric indica- tors of maturation for populations of L. sinuensis. Resulting implications for management of penaeid shrimp ponds are discussed. Materials and methods All animals were obtained at the Agrosoledad S.A. shrimp farm on the upper Cispata estuary of the Rio Sinii, Departamento de Cordoba, Colombia, near 9°17'N, 75°50'W. In order to avoid overt effects of pond harvesting, feeding, and other management activ- ity, collections were restricted to bottom sediments of earthen drainage canals that extend for several kilometers through the farm. Specimens were indi- vidually extracted from burrows with yabby pumps (Manning, 1975), and samples included all animals retained when extracted sediments were washed on coarse (5-mm) sieves. All population samples were taken within 500 m of the upper end (origin) of the drainage canal system where colonized bottom muds ranged from about 3 to 6 m in width and population density ranged from 15 to 40 animals per m-. Monthly sampling of populations extended over four years from December 1991 through December 1995. Each monthly sample consisted of animals obtained from burrow openings encountered at ran- dom during walks along transects that crossed ex- posed mudflats at the sampling site during a period of reduced pond discharge. The sample consisted of at least 60 animals, and over 3200 animals were col- lected in the course of the study. Following collec- tion, animals were transported in individual perfo- rated vials (Felder, 1978) to the farm laboratory for morphological analysis. Specimens lacking chelipeds. having incompletely regenerated chelae, or with other obvious appendage deformities were excluded from the analyses that involved measures of those appendages but included in analyses of population size-class structure. In the laboratory, each individual was sexed on the basis of anterior pleopod morpho- logical features and evidence of ovaries visible through the integument. Wet weight (WW) was de- termined to ±0.1 g on a top-loading balance after animals were blotted with tissue paper. Dry weight was estimated by drying the animals at 60°C for 48 h, and ash content was estimated from weight after ignition at 500"C for 6 h. When present, eggs on pleo- pods were included in weights of ovigerous females. Morphometric measurements (Felder and Lovett, 1989) were made with dial calipers to ±0.05 mm. Because carapace length (CD is usually less depen- dent on gonadal development than are most other size measurements (Hartnoll, 1982; Felder and Lovett, 1989), it was selected to represent body size and measured from the tip of the rostrum to the pos- terior margin of the cardiac region. Total length (TL) was measured from the tip of the rostrum to the pos- terior margin of the extended telson. Major chela height ( ChH ) was measured as the maximum height of the propodus, inferior margin to superior margin. Major chela width (ChW) was measured as the maxi- mum width of the propodus from the most convex area on the internal surface to the opposite convex area on the external surface. Major chela length (ChL) was measured as the maximum length from the superior proximal articulation of the propodus with the carpus to the superior distal articulation with the dactylus. Between November 1992 and De- cember 1995, ovarian width (OW) of females (the width of the right ovary visible dorsally through the integument of the third abdominal segment) was determined. An index of relative ovarian development was estimated from the ratio OW/CL. Color and num- ber of eggs, color of the ovaries, evidence of para- sites, and occurrence of damaged or missing chelae were also recorded. Egg size was measured under a light microscope and ocular micrometer. Volume of the eggs was calculated from the mean of the long and short axes used as the single diameter measure- ment for a sphere (McEdward and Chia, 1991). Temperature and salinity were measured twice daily (dawn and dusk) at the surface and bottom of filled ponds. Temperature was measured ±1 °C with the probe of a YSI® model 57 temperature-compen- sated oxygen meter. Salinity ± 1 ppt was measured with a temperature-compensated refractometer. All means were reported along with the 95% CI (confidence interval). Monthly population samples were pooled to compare quarterly variations in the 528 Fishery Bulletin 97(3), 1999 distribution of size cohorts and the occurrence of ovigerous females in the population. Temporal trends in both carapace lengths and sex ratios for the over- all four-year span of the study were analyzed by lin- ear-linear piecewise polynomial regression to define slope transition points. Percentile data for the sex ratios were arcsine-transformed prior to these analy- ses. In order to remove effects of morphologically anomalous postreproductive and senescent individu- als, animals >18.0mm CL (two females and three males) were excluded from analyses of relative growth. Linear-linear piecewise polynomial analy- sis of untransformed data was used to optimize posi- tioning of ontogenetic transition points (±0.05 mm) in all relative growth comparisons. Except for this alternative method of iteratively locating slope tran- sitions, methods complied with the recommended practice of fitting regi'essions to untransformed data above and below transition points by reduced major axis (Lovett and Felder, 1989). Standard allometric coefficients (Huxley, 1932 ) were also determined from regressions of log- transformed data by least squares estimate, with data subdivided at the previously es- timated transition points. All statistical analyses were performed with NCSS® (Number Cruncher Statistical System) 6.0 software (Hintz, 1995). Because generic revisions have yet to address many of the confamilial taxa that are treated in our com- parative discussions, it was necessary in compara- tive discussions to refer to some species under Callianassa sensu lato (s.l.), while acknowledging that Callianassa sensu stricto has been restricted to a few eastern Atlantic populations (see Manning and Felder, 1991). Results Sex ratio Except for January to June 1992, March 1993, and February 1995, sex ratios were significantly female- biased over the four years of sampling (Fig. lA). Over the entire study period, the mean sex ratio was 2.4 +0.3 females per male, with the lowest mean ratio (1.1) occurring in February 1992, and the highest ratio (5.1) in October 1993. Trends in mean CL of the population (Fig. IB) generally tracked those in sex ratios, as defined by piecewise regression analy- sis. Slopes over the first two years revealed declin- ing relative abundance and increasing mean CL of males, patterns that differed significantly from those of the last two years of monitoring. An iteratively optimized transition point for sex ratios occurred at 23 months, whereafter mean sex ratios remained near asymptotic. Analysis over the full study period also defined an optimized transition point or peak in CL of males at 23 months, beyond which male CL was asymptotic. The CL of females, however, peaked late in the first year of monitoring. Mean size of both males and females increased slightly but significantly from the first through fourth quarter of each year except for 1994, a year in which the presence of first quarter recruits was offset by a cohort of large individuals surviving from the previ- ous year (Fig. 2). When treated separately, mature- sized individuals (>11 mm CL) of both male and fe- male populations increased in mean CL over the year, usually peaking in the fourth quarter prior to maxi- mum abundance of ovigerous females in the follow- ing year (Fig. 3, B and D). Mean CL of males ap- peared to exceed slightly that of females over the full study period, and males dominated large size classes (Figs. IB, 2, 3B). Only 4.37f of the collected females were >15.0 mm CL, whereas 12.4% of males attained this size. The largest animal collected was a 20.7- mm-CL male. Detection of recruits <8-10 mm CL was sporadic throughout the study (Fig. 3E), and tracking of growth in recruitment cohorts was not possible. Ovigerous females Among all ovigerous females collected during the study (/!=444), size ranged from 7.04 mm CL to 16.8 mm CL. However, 38.3'7( of all these were in the 13.0 to 13.9 mm CL size class and <1.59^ of ovigerous fe- males were <11.0 mm CL (Fig. 4). The single oviger- ous specimen <9.8 mm CL was an apparently preco- cious individual taken in August 1994, while the population of ovigerous females was low (Fig. 3D). During peak abundances of ovigerous females (Feb- ruary to June, Fig. 5B), 64.3'/f of females in the 13.0 to 13.9 mm CL size class were ovigerous (Fig. 2). Ovigerous females were found in all months of the four-year study (Figs. 2 and 3D). However, percent- ages were lowest in the third quarters of 1992, 1993, and 1995, and the fourth quarter of 1994, whereas the highest percentages occurred in the first or sec- ond quarter of each year. Peak abundance of oviger- ous females coincided with the highest quarterly sa- hnity in 1992, 1994, and 1995 but was less defined in 1993 when the fourth quarter salinity did not fall toa typical low value (Fig. 3, A and D).When monthly samples were pooled over the entire study, highest percentages of ovigerous individuals among all fe- males were evident during the high salinity period from February to June; a maximum was found in May (35.7%), three months after peak development of the ovaries (Fig. 5, A-C). Nates and Felder: Growth and maturation of Lepldophthalmus sinuensis 529 B B so a. s. a U 95 90 85 80 75- 70- 65- 60- 55 50- 45- 16- 15- 14- 13- 12- 11- 10- I I I I I I I I I I I I I M I I I I I I I I I I I I II I I I I I I I M I I I I I I I ■5 ■10 ■15 ■20 ■25 SL -30 S ■35 ■40 -45 -50 ■55 9 I II I I I I I I I I I I I I I I I I II I I I I I I I I I I I I I II I I I I I I I I I I M I I Fb My Ag Nv Fb My Ag Nv Fb My Ag Nv Fb My Ag Nv 1992 1993 1994 1995 Figure 1 Sex composition (A) and mean carapace length iCLi (B) in sampled Colombian populations of Lepldophthalmus smuensis. December 1991 through December 1995. Mean CL is shown separately for males (closed circles) and females (open circles); vertical lines define 95'r CI. The transition point at which data are subdivided, estimated by piecewise linear-linear polynomial regression, was selected to minimize the sum of squares of residuals. Development of ovaries reached maxima in the quarter prior to peak occurrences of ovigerous fe- males and either coincided with or immediately fol- lowed the quarter in which lowest ambient salini- ties occurred (Fig. 3, A, C, and D). Examined as a monthly value pooled over the three years that it was monitored (Fig. 5C), this index of ovarian develop- ment was at minimum values in June and July when ovaries of many spent and immature females, al- though evident for the full length of the abdomen, were limited in width to narrow translucent-yellow strands. Ovaries became deep yellow to yellow-or- ange as relative ovarian width began to increase markedly in August, concurrent with an annual de- cline in ambient salinity. By October, ovaries of most mature females were somewhat lobate in shape and yellow-orange to reddish orange in color, and ova- rian index reached high values that persisted through 530 Fishery Bulletin 97(3), 1999 ^m ^ t~W trx im. d rf: i: "i; '" 'hm r i: (%) Xouanbsjj J3 s S 6 3 ,0' CO m II 3 C o — .2> ll > -H s ^ 2 -^ - c — Qj «) o ^Q s^-a D- 3 = t; ■« -fi CC 01 H O^ 3 ^ -2 •■^' - 1 3 = CT g th a ^ -c J •« O a — o en "a ^& N O CO w >■§ I- ■ ri 3 S- February. Highest values for the ovarian index, in Febru- ary (0.139 ±0.012), were sig- nificantly greater than were preceding values in Decem- ber or those that followed in March through May. De- creasing measures of relative ovarian width during April through June (Figs. 3C and 5C ) were coincident with the second quarter maxima in abundance of ovigerous fe- males (Fig. 3D and 5B) be- cause many of those females had spent ovaries. Shape of the eggs was al- most spherical in recently deposited pale yellow to yel- low-orange clutches and be- came more oblong as color faded to translucent gray- brown in late-stage eggs just prior to hatching. Mean egg volume in live late stage eggs was 0.81 ±0.07 mm^ (/! = 76 eggs, from 5 different fe- males), and the coefficient of variation in volume within a single egg clutch was 5.09% (/;=30); mean dimensions were 1.22 ±0.04 mm by 1.05 ±0.02 mm. The total number of eggs per female ranged from 182 ±44 in small fe- males (<12.0 mm CL; «=30) to 301 ±36 in large females (<14.0 mm CL; /7 = 141). The overall mean number of eggs for all ovigerous females col- lected was 251 ±18 (Ai=:444), and the maximum number found on any single female was 958 on a specimen of 14.2 mm CL. Monthly egg counts per ovigerous female, pooled over the study (Fig. 5D), re- vealed peak abundances in February-April, followed by significantly lower numbers in June-July. Thus, the over- all rate of egg production, when considered as a prod- uct of both abundance of ovigerous females and egg Nates and Felder: Growth and maturation of Lepidophthalmus stnuensis 531 abundance per clutch (Fig. 5, B and D), was maintained at par- ticularly elevated levels from February through at least April and perhaps May. Growth and sexual maturation 30 10 0 £ 14 G 12^ 0 B ~ 0.16 E E - c u E E o 3 E 0.12- 0.08 Size at sexual maturation in both males and females of the sampled population was evident in rela- tive growth changes of the major chela. Asymmetry of the first pereopodal chelae was clearly evident in both sexes and did not show bias for right or left hand- edness in random subsamples (left:right handedness 23:27 for males, 25:25 for females; /i=50 per sex). The value of this feature as a secondary sex character was suggested initially by the much heavier general appearance of the appendage in large males than in females of approximately the same size, and sexual differ- ences in allometric growth were documented subsequently by re- gression analyses of morphologi- cal measurements ( Fig. 6; Tables 1 and 2). Relative growth rate in the chela width of males increased in animals that had reached a mean carapace length of at least 11.3 mm, as estimated by linear-lin- ear piecewise polynomial regres- sion analysis (Fig. 6A). Indepen- dent regression analysis of males larger and smaller than this size by reduced major axis revealed significant differences in slopes for those data sets ( Table 1 ), even though the two values for this positive allometric growth did not differ significantly when com- pared as allometric coefficients (Table 2 ). When fitted by reduced major axis, regression slopes for both prematuration and postma- turation phases of growth in female chela width (Fig. 6B) differed significantly from those for both growth phases in males, although there was far greater dif- ference between the sexes in the postmaturation phase. Relative growth in the chela width of females Figure 3 Quarterly measures of salinity, growth, and reproductive indices in sampled Co- lombian populations of Lepidophthalmus sinuensis, January 1992 through De- cember 199.5; for A-C, vertical lines define 95% CI; for D-E, vertical lines define range. (Al Morning surface salinity Ippti at the farm intake pump station in the upper estuary, e.xpressed as mean of daily measures throughout each quarter; (B) mean CL of mature Oil mm CLI males (closed circles) and females (open circles) in sampled populations; (C) relative ovarian development (OW/CL); (D) percent- age of ovigerous individuals among females; (E) percentage of juveniles (<10.00 mm CLi. became significantly less positive beyond a carapace length near 10.8 mm. The negative allometric growth (coefficient <1 ) for this postmaturation phase in females (Table 2) differed significantly from the strongly posi- tive allometric growth in postmaturation males. 532 Fishery Bulletin 97(3), 1999 40 30- 20 10- Analysis of relative growth in ma- jor chela height for males ( Fig. 6C ) and females (Fig. 6D), when scaled to cara- pace length, produced remarkably similar results to that for relative growth in chela width, suggesting some interdependency of those mea- sures. Transition points were opti- mally located at a mean carapace length of 11.0 mm for males and 11.2 mm for females, and postmaturation growth differed significantly between the sexes (Table 1). On the basis of al- lometric coefficients (Table 2 ), positive prematuration allometric growth once again did not differ between the sexes, whereas negative postmaturation growth in females differed signifi- cantly from the strongly positive allo- metric growth in postmaturation males. Analysis of chela width scaled to chela height in both sexes did not indicate signifi- cant prematuration differences between males and females, and growth in both sexes was isometric or nearly so (Tables 1 and 2). However, following tran- sition to the postmaturation phase in males, rela- tive growth in chela width exceeded that of height. The opposite effect was seen in postmaturation fe- males; negative allometric growth was seen in chela width, when compared to chela height. Analysis of total length scaled to carapace length revealed large measures of error (Tables 1 and 2), as expected in size measurements that include the length of a soft, stretchable abdomen. Allometric co- efficients indicated that slopes did not significantly differentiate prematuration growth from an isomet- ric pattern in either sex, whereas postmaturation growth on the basis of this parameter appeared to be negatively allometric but similar in rate for both males and females. Large measures of error associated with wet weight measurements in relation to carapace length, espe- cially at small sizes, limited detection of possible sex differences in allometric coefficients based upon this parameter and did not resolve significant differences between growth phases of either sex (Table 2). Only in the postmaturation phase were allometric coeffi- cients based upon wet weight significantly more posi- tive in males than in females. The more sensitive analyses of wet weight based upon regressions by reduced major axis (Fig. 7, A and B; Table 1) detected highly significant differences between postma- turation males and females, regardless of whether ovigerous females were included or excluded from the analysis, and also resolved significant differences V//J//J/A 10 11 12 13 14 Carapace length (mm) 15 16 Figure 4 Size class (CL) frequency ciistribution of all ovigerous females in sampled Colombian populations of Lepidophthatmus sinuensis, December 1991 through December 1995 (number by class appears above each bar I. in rates of mass accumulation between prema- turation and postmaturation growth phases. Opti- mized transition points in piecewise regression analy- ses of wet weight to length relationships were near 9.8 mm CL for females and near 11.3 mm CL for males. These regression breaks suggest that matu- ration in males occurs at slightly higher wet weights ( 1.5 to 2 g) than in females (usually near 1 g). Dry weights and ash free dry weights were deter- mined for limited subsets of our sample and were not subjected to regression analysis. The proportional relation of dry to wet weight in ovigerous females (0.47 ±0.06; /!=8) was slightly but significantly higher than in nonovigerous females (0.36 ±0.03; n=76} and males (0.33 ±0.05; n = l2), whereas the relation of ash free dry weight to wet weight in ovigerous females (0.052 ±0.003; n=8) was not significantly different from that of either nonovigerous females (0.059 ±0.011; n=41) or males (0.055 ± 0.003; n=2). Discussion Sex ratio Strongly female-biased sex ratios in the monitored population of Lepidophthalmus sinuensis are in marked contrast to those reported in intertidal popu- lations of L. louisianensis on the Mississippi coast in the northern Gulf of Mexico (Felder and Griffis, 1994), which were variable but averaged near 1:1. However, female-biased ratios have been reported in populations of L. louisianensis from the Louisiana coast in the northern Gulf of Mexico (Felder and Nates and Felder: Growth and maturation of Lepidophthalmus stnuensis 533 35 30 25- 20 15H 10 5 0 E E^ _i u f O Lovett, 1989), as well as in populations of western Pacific intertidal thalassinideans such as Trypaea australiensis Dana (Hailstone and Stephenson, 1961), Callia?}assa s. 1. filhoU Milne Edwards (Devine, 1966). and sexually mature adult popu- lations of Neotrypaea califor- niensis (Dana) and Upogeblci pugettensis Dana (Dumbauld et al., 1996). Populations were male-biased in sampled popula- tions of Callianassa s. 1. subter- ranea Montagu from subtidal European waters (Rowden and Jones, 1994). Observance of skewed sex ra- tios in thalassinidean samples may reflect actual bias in the population or an artifact of col- lecting methods that do not sample sexes equally within the vertical profile of burrows or across horizontal beach transections, especially when collections must be restricted to readily accessed reaches of shoreline sediments. Move- ment of ovigerous females to- ward the surface for egg release or to optimally ventilate eggs could lead to greater probabil- ity of their capture when bur- rows are aspirated with yabby pumps. Likewise, sampling re- stricted to shorelines may re- flect differential positioning of sexes within the burrow along tidal or wave agitation clines. Such effects could account for previous observations in which populations of Callichirus islagrande were female-biased on high wave-energy beaches but not on nearby protected beaches, or for ratios that dif- fered markedly from month to month in some populations of L. loiiisianensis from the Gulf of Mexico (Felder and Griffis, 1994). In the present study, mudflats along margins of drainage canals were fully accessible to sampling during periods of low pond discharge, samples were not limited to shorelines, and burrows were not as deep as reported E Figure 5 Pooled monthly measures for salinity, temperature, and reproductive indices in sampled Colombian populations of Lepidophthalmus stnuensis: for A, monthly tem- perature data were pooled from December 1991 through December 1993, and monthly salinity from January 1992 through December 1995; for B, monthly data were pooled from January 1992 through December 1995; for C, monthly data were pooled from January 1993 through December 1995; for A, C. and D, vertical lines define 95Tf CI; for B, vertical lines define range. (A) Morning surface salinity i solid circles, continu- ous line I at farm intake pump in upper estuary, and afternoon surface temperature (open circles, broken line) at outlet in pond 1. expressed as means of daily measures; (B) frequency of ovigerous individuals among mature oll.O mm CL) females of populations; lC> relative ovarian development (OW/CLi; iD) mean number of eggs in clutches of ovigerous females. for above mentioned populations of C islagrande and L. loiiisianensis. From resin casts (Fig. 4 in Nates and Felder, 1998), it was observed that almost all burrows were <1 m in depth, thus encompassing water volumes readily extractable with yabby pumps. 534 Fishery Bulletin 97(3), 1999 Table 1 Formulae and statistics for linear regressions (by reduced major axis, with untransformed data) of chela width on carapace length lChW:CL). chela height on carapace ler gth (ChH:CL). chela width on chela height (ChW:ChH), total length on carapace length (TL:CL) and wet weight on carapace ength (WW:CL) for male and female populations of Lep idophthalmus sinuensis sampled at the Agrosoledad S. A. shrimp farm upper Cispata estuary, Colombia, from December 1991 through December 1995. Regression s were calcul ated for entire data set I All ) and for data subdivided at the transition point ( yielding separate regressions for data po nts X where A' = size at onset of sexual maturity estimated by piecewise linear-linear polynomial | regression) > 'i = sample size, r- = coefficient of determination, 95% CI = confidence interval of slope. n r2 Formulae 95% CI ChW:CL Males All 985 0.69 C/iW = (0.612 xCL) - 3.474° ±0.021 X 767 0.49 ChW = (0.789 yCL) - 5.974°'' ±0.039 Females All 1998 0.35 CAW = (0.273 xCD- 0.519°'' ±0.009 X 187.3 0.12 ChW = (0.304 xCL) - 0.943°'' ±0.013 ChH:CL _. Males All 987 0.74 ChH = (0.991 ^CD- 4.843' ±0.031 X 795 0.57 ChH = (1.198 xCD- 7.76r ±0.054 Females All 1998 0.49 ChH = (0.513 xCL) - 0.425'^'' ±0.016 X 1821 0.20 CA/f = (0.513 xCD- 0.425'^'' ±0.021 ChW:ChH Males All 987 0.94 ChW = (0.618 xChH) - 0.482''/' ±0.009 x 713 0.88 ChW = (0.655 xChH) - 0.83r« ±0.017 Females All 2000 0.71 ChW = (0.532 -.ChH) - 0.292''« ±0.013 x 1929 0.53 ChW = (0.588 xChH) - 0.646' ±0.018 TL:CL Males All 1007 0.80 TL = (4.349 xCZ.1- 1.033'"' ±0.121 X 270 0.15 TL = (6.374 xCL) - 32.918*^ ±0.703 Females All 2158 0.64 TL = (4.641 xCL)- 3.186/-'* ±0.117 SX 1006 0.64 TL = (5.419 xCL) -11.49,5'"-' ±0.201 >X 1152 0.22 TL = (5.766 xCZ.) -19.813* ±0.293 WW:CL Males All 937 0.70 WW = (0.751 xCL) -6.431''"" ±0.026 X 737 0.60 WW = (0.998 xCL) - 9.927'"'» ±0.046 Females incl uding th ose with eggs) All 1933 0.49 WW = (0.613 xCL) - 5.0.53'P ±0.019 x 1883 0.42 WW = (0.698 xCL) - 6.204''"' ±0.024 Females excl uding th ose with eggs) All 1542 ■ 0.53 WW= (0.581 xCL) - 4.680" ±0.019 X 1501 0.44 WW = (0.662 xCL) - 5.767°'-» ±0.024 as = allometric coefficient; with same superscript are significantly (f <0.05 different from each other. Nates and Felder: Growth and maturation of Lepidophthalmus sinuensis 535 Table 2 Allometric coefficients measured as slopes for linear regressions I by least squares estimate, with log-transformed data) of chela width on carapace length (ChW:CL), chela height on carapace length (ChH:CLi. chela width on chela height lChW:ChH). total length on carapace length (TL:CL), and wet weight on carapace length (WW:CL) for male and female populations of Lepidophthalmus sinuensis sampled at the Agrosoledad S. A. shrimp farm, upper Cispata estuary, Colombia, December 1991, through December 1995. Coefficients were determined for entire data set (All) and for data subdivided at the transition point (yielding separate regressions for data points X, where X = size at onset of sexual maturity estimated by piecewise linear-linear polynomial regression), 95'7f CI = confidence interval of coefficient. * *= allometric coefficient significantly (P<0.05) different from 1.0. a-u = allometric coefficients with same superscript are significantly (P<0.05) different from each other ChW:CL ChH:CL ChW:ChH TL:CL WW:CL Allom. coef. 95% CI Allom. coef 95% CI Allom. coef 95% CI Allom. coef 95% CI Allom. coef 95% CI Males All 1.577**° X 1.608**' ±0.059 ±0.162 ±0.129 1.457**X 0.441**'"*''' ±0.053 0.491***'/*'' ±0.047 ±0.122 ±0.110 ±0.033 ±0.160 ±0.045 1.085 1.061 1.106 0.934'*'-'*' 1.096' 0.864**'-'*' ±0.017 ±0.039 ±0.033 ±0.021 ±0.116 ±0.033 0.928**"'' o.ges"'' 0.618**'""° 0.895**''"'' LOSl"""" 0.693**'"'' ±0.027 ±0.035 ±0.164 ±0.025 ±0.052 ±0.051 3.094**" 2.762** 3.056**' 2.476**'" 3.458**" 2.064**'''" tO.118 to. 747 ±0.118 tO.090 tl.102 ±0.122 This observation suggests that the detected sex bias accurately characterizes the population sampled. In the course of previously reported observations on populations of C. s. 1. filholi, it was noted that the sex ratio was near 1:1 only when small animals were included in the sample (Devine, 1966). For L. sinuensis there is also a correlation between mean size of animals in the monitored population and sex ratio over the four years of monitoring. As noted for C. s. 1. filholi, samples of smaller mean CL were nearer 1:1 in sex ratio than were samples of larger mean CL. The trend toward increasing frequency of females reached an approximate asymptote after 23 months of observation, simultaneous with the trend in increasing CL of males, and may indicate a point at which increasing numbers of larger males estab- lished a sexual equilibrium by displacing small males from a limited habitat or maintained territory. This increase corresponds roughly to the time when bur- row densities within ponds on the farm began expo- nential increase and when detrimental effects on water quality and penaeid production became evi- dent (equals month 38 in Figure 1 of Nates and Felder, 1998). Sex ratios and mean carapace lengths may thus serve to forecast potential for rapid in- creases in infesting populations. Both the large body size and the robust chelipeds of mature males could be of advantage in competi- tion for limited space and available females wher- ever L. sinuensis occurs in dense, mature aggrega- tions. Such behaviors have been suggested previously to account for large numbers of possibly displaced juvenile males of L. louisianensis observed periodi- cally in plankton samples (Felder and Lovett, 1989; Felder and Rodrigues, 1993). Similar intraspecific competition for space, in which juveniles are expelled to the surface, has also been postulated to occur in some populations of Callianassa s. 1. japonica ( Ortmann ), along with possible immigration of juve- niles to other habitats (Tamaki and Ingole, 1993). While not documented to occur in L. sinuensis, colo- nization of mariculture ponds by such displaced ju- venile males pumped in with estuarine waters could account for higher percentages of males observed early in our study. Such recruitment of juveniles could potentially supplement larval settlement within ponds to rapidly build early infestation densities. Ovjgerous females The abbreviated life cycle in L. sinuensis includes only two brief zoeal stages, large numbers of which occur in nocturnal plankton samples taken at the study site (Nates et al., 1997). The eggs of this spe- cies and its congener L. louisianensis are comparable in size to those of Callianassa s. 1. kraussi Stebbing, another estuarine species that exhibits markedly abbreviated development (Forbes, 1973), and are larger than those of many thalassinids with longer larval histories and wider planktonic dispersal (see 536 Fishery Bulletin 97(3), 1999 Pohl, 1946; Forbes, 1973; Tunberg, 1986), with some clear exceptions (see de Vaugelas et al., 1986). Given abbreviated development in L. sinuensis and the like- lihood of almost immediate settlement by decapodids into the vicinity of our sample site, we expected to detect distinct cohorts of small recruits, from which we could estimate rates of growth to maturation. However, unlike the population of the warm temper- 3 4 U Males 0 13 11 I 9 Females B Males Females 9 12 15 18 21 3 6 9 Carapace length (mm) 12 15 18 21 Figure 6 Linear regression (by least squares estimate for untransformed data) of chela width (ChW) on carapace length (CL) for (A) males and (B) females and chela height (ChH) on CL for (C) males' and (D) females in sampled Colombian populations of Lepidophthalmux sinuensis. December 1991 through Decem- ber 1995. The transition point at which data are subdivided, estimated by piecewise linear-linear polynomial regression, was positioned to minimize the sum of squares of residuals. ate species L. louisianensis from the Gulf of Mexico (Felder and Lovett, 1989), tropical populations of L. sinuensis included ovigerous females in every month of sampling. This potential for continuous recruit- ment over the annual cycle, coupled with our lim- ited success in consistently capturing small individu- als, limits clear definition of reproduction and growth cycles from collections of recruits. Vitellogenesis in L. sinuensis was evident in coloration and size changes of ovaries viewed through a translu- cent region of the cuticle, as in other thalassinid species (Hailstone and Stephenson, 1961; Forbes, 1977; Felder and Lovett, 1989). The pattern of change in L. sinuensis over the course of maturation is as much as that reported for L. louisianensis, with ovaries becoming more massive and opaque as maturation advances (Felder and Lovett, 1989). However, when tracked as an annual index of reproductive activity, relative ovarian width in L. sinuensis at no point reached the highest mean values re- ported for L. louisianensis (even af- ter correcting misplaced decimals in the _v-axis of Fig. 1 of Felder and Lovett [1989]). We suggest that this reflects year-round reproductive ac- tivity in L. sinuensis, rather than in- vestment in a more punctuated tem- perature-modulated event as might be expected in temperate latitudes. It may also account for a greater number of eggs on ovigerous females of L. louisianensis (598 ±212; n=4) than on ovigerous females of the year- round reproducing L. sinuensis (251 + 18; «=444) (Nates et al., 1997). Even so, significant fluctuations in mean values for relative ovarian width did occur over the three years that we monitored this value in L. sinuensis, with highest values preceding maxi- mum abundance of ovigerous females by two to three months. In tropical nearshore decapods, it is common to have extended periods of reproduction without distinct peaks defined by seasonal tempera- ture (Sastry, 1983; Steele, 1988; Bauer, 1992; Mouton and Felder, 1995). However, whether modulated by temperature or other factors, an- D Nates and Felder: Growth and maturation of Lepidophthalmus sinuensis 537 nual cycles in availability of nu- trition may determine potential for vitellogenesis and success of larvae (Zimmerman and Felder, 1991: Mouton and Felder, 1995). Although temperature was not subject to marked fluctuation in the shrimp farm habitat of L. sinuensis, it did vary slightly with the annual changes in rain- fall patterns that determined sa- linity of the upper estuary (Fig. 4A). The annual decline in salin- ity from April through October reflected increased input of nu- trient-loaded waters from the nearby Rio Sinii which was re- ported to result in periodic eleva- tion of penaeid shrimp growth rates on the Agrosoledad farm (MogollonM. Similarly, such low salinity periods may have pro- duced improved conditions for vi- tellogenesis inL. sinuensis. Peak development of ovaries in the first quarter of the typical year either coincided with or shortly followed lowest mean salinity during the last quarter of the pre- vious year, and fell markedly just after salinity peaked. The abundance of ovigerous females peaked almost simultaneously with salinity in most years of the study and a decrease in mean ovarian width im- mediately followed, indicating that the abundance of females with spent ovaries was not offset by other females simultaneously undergoing vigorous ovarian development. Reduced vitellogenesis following the high salinity period was also suggested by the mark- edly reduced mean number of eggs per clutch found in those ovigerous females that occurred from May through July. Growth and sexual maturation Among callianassids and other thalassinid shrimp, the major chela is used commonly for aggressive in- teractions between captured individuals that are held together or among individuals encountering one an- other where burrows intersect in laboratory fossoria (MacGinitie, 1934; Pearse, 1945; Rodrigues, 1976; Tunberg, 1986; Felder and Lovett, 1989; Pillai, 1990; 20 4 8 Carapace length (mm) Figure 7 Linear regression ( by least squares estimate for untransformed data) of wet weight (W^\') on carapace length (CD for males (A) and females (B) in sampled Colom- bian populations o( Lepidophthalmus sinuensis, December 1991 through Decem- ber 1995. The transition point at which data are subdivided, estimated by piece- wise linear-linear polynomial regression, was positioned to minimize the sum of squares of residuals. ' Mogollon, J. V. 1995. Agrosoledad S. A., Playa de la Artillena no. 33-36. Cartagena. Columbia. Personal comniun Rowden and Jones, 1994). Contrary to the case with many other genera, both males and females of Lepi- dophthalmus have one chela enlarged, but the ma- jor chela of mature males is much more strikingly developed than that of mature females (Felder and Rodrigues, 1993; Felder and Manning, 1997). Sexual dimorphism of the chelae in decapods is generally thought to reflect adaptations for their use. especially by males, in combat, display, and courtship (Hartnoll, 1974). The enlarged chela can play a role in intermale competition during the precopulatory phase of the life cycle, and marked change in its allometric growth rate is frequently related to functional sexual matu- rity (Grey, 1979; Aiken and Waddy, 1989; Claxton and Govind, 1994; Gu et al., 1994; Robertson and Kruger, 1994). This is likely also the case in L. sinuensis, even though we can only speculate as to the exact role this appendage plays in such fossorial populations. A striking transition takes place in rate of relative growth for the major chela in L. sinuensis. and there is an allometric divergence between males and fe- males in the shape of this appendage at sizes that exceed a defined, transitional carapace length. For the most part, the relationships of slopes above and below these transition or maturation points are simi- lar to relationships previously reported in monitored 538 Fishery Bulletin 97(3), 1999 populations of L. louisiarjensis from the northern Gulf of Mexico (Felder and Lovett, 1989). In relative growth parameters of the major chela, when scaled to carapace length, males and females of both Lepidophthalmus species are much more similar to one another at sizes below transition points than at sizes exceeding those points. The estimated maturational transition size in male L. sinuensis ranges from 11.0 to 11.3 mm CL, whether based upon regression analysis of relative gi'owth in chela width, chela height, or wet weight. The range of these values is markedly less than that reported in L. louisianensis (>15.5 mm CL), a warm-temper- ate congener (Felder and Lovett, 1989). In all of these measures, postmaturation growth rate in males is more positively allometric and significantly greater than postmaturation growth rate in females, as has been shown in many other decapods (Hartnoll, 1978, 1982 ). Although relative weight increases were posi- tively allometric in both males and females of ma- ture size, the positive allometry of growth in size of the major chela, accompanied by sex-dependent changes in its shape, appears to account for postma- turation rates of weight increase in males that sig- nificantly exceeding those of females. Weight ac- cumulation due to massive but episodic development of ovaries in maturing females would appear to be somewhat offset by the simultaneous trend toward negative allometric growth of the major chela. Estimated from the major chela measurements, the mean maturational size in female L. sinuensis ranges from 10.8 to 11.2 mm CL, or slightly less than that of males but very near the values reported for females of L. louisianensis, which ranged from 10.7 to 11.0 mm CL ( Felder and Lovett, 1989). Because over 99% of the ovigerous females we collected were >11.0 mm CL, the general validity of this mean size estimate for maturation appears to be confirmed on an inde- pendent basis, even though two atypically small ovigerous females <10.0 mm CL, were collected (Fig. 4). Our regression analysis of wet weight indicated that the transitional maturation point for females fell near 9.8 mm CL, but error appeared to be high in values surrounding this point. A relatively large range and error in this weight-dependent value should perhaps be expected because two variables, negative allometry of the major chela and growth of the ovaries, are likely influencing it in opposite ways. We suggest that the aforementioned annual salinity variations and the simultaneous effects on nutrient loading may seasonally alter rates of ovarian growth. This could in turn contribute to the range of weight dependent values for maturation or account for oc- casional occurrences of small egg-bearing females by favoring early maturation of young cohorts. Management summary Those penaeid culture ponds on the Caribbean coast of Colombia that operate at higher salinities than those near the Rio Sinu have not been found to har- bor detrimental infestations of Lepidophthalmus. However, nutrient-rich, estuarine waters are believed to yield high rates of penaeid production, promoting location of several major penaeid farms in oligohaline habitats that harbor natural populations of L. sinuensis. This pest is introduced to culture ponds by pumping in larvae and perhaps migrant juveniles from the estuary. Although filtration of influent wa- ter should serve to limit entry of any migrant juve- niles, it is practical to limit entry of much smaller planktonic lai-vae only by carefully avoiding pump- ing in of waters from the estuary during nocturnal hours of peak larval activity in the water column ( Nates et al. , 1997 ). Because we herein document that L. sinuensis reproduces year-round, this diel restric- tion of pumping schedules cannot be limited solely to the first annual quarter of elevated salinity when peak abundances of ovigerous females suggest that high densities of larvae might be present among noc- turnal plankton. However, such periods would almost certainly present the greatest risk of larval intro- ductions to culture ponds. Although controlling water influx may limit colo- nization of ponds from surrounding environments, Lepidophthalmus will almost certainly over time succeed in becoming established in estuarine ponds that provide it with both richly organic muddy sub- strates and annual cycles of water quality that favor reproduction. Detrimental effects become obvious when densities in ponds exceed 200 burrow openings/m-^ or 66 animals/m- (Nates and Felder, 1998), levels we have never found in natural settings on in the Rio Sinii estuary. Explosive reproduction within the ponds appears to immediately precede problem levels of infestation, and measurements of the sex ratios and CL of the population may serve to forecast the po- tential for such an event. Because attainment of a strongly female-biased sex ratio is roughly coincident with previously reported exponential population growth (Nates and Felder, 1998), both sex ratio and density of mature females may prove useful predic- tors of population growth potential. Maturity of both males and females can be determined by measures of CL and cheliped dimensions, or by inference from burrow diameters, and immediately pending episodes of egg deposition can be predicted by monitoring the ovarian index. Dense infestations of Lepidophthalmus in Colom- bian shrimp culture ponds have to date been treated periodically by farm operators with the carbaryl pes- Nates and Felder: Growth and maturation of Lepidophthalmus sinuensis 539 ticide Sevin®, a practice unlikely to change in the absence of proven alternative management strate- gies. From our observations, such applications have usually been contained within drained or slightly filled culture ponds and limited only to control of ex- cessive thalassinid densities. It is possible that re- ductions in frequencies of such applications can be achieved by timing of treatments to pre-empt repro- ductive peaks or by limiting pesticide applications only to infestations that are found to include high percentages of mature or near-mature females, such as we observed in this study most often in the fourth annual quarter. Any measure that reduces applica- tion rates and frequencies may lessen ecological risks and contribute to effectiveness of carbaryl degrada- tion prior to flushing treated ponds into surround- ing environments. Acknowledgments We thank G. T. Rizzuto, R. G. Jaeger, R L. Klerks, and R. R. Twilley, University of Southwestern Loui- siana, and R. Lemaitre, Smithsonian Institution, for comments on earlier versions of the manuscript. We are also grateful to many associates of the Agro- soledad S. A. shrimp farm, particularly J. V. Mogollon, L. Mogollon, G. Moreno, and M. Torralvo, who provided access to sites in Colombia, and A. Ar- royo, D. Espitia, and S. Garcia, who assisted in col- lecting population data there. Among others who assisted with field and laboratory work, we thank L. Borda, R. Bourgeois, V. Fuentes, R. Griffis, R. Lemaitre, C. Moreau, S. de A. Rodrigues. and K. Strasser. Fellowship support for this project was pro- vided to S. F. Nates from the Institute Colombiano para el Desarrollo de la Ciencia y la Tecnologia Fran- cisco Jose de Caldas (COLCIENCIAS), Colombia. Logistical support was provided through ACUANAL, BANCOLDEX, and the efforts of C. M. Sanin. Sup- port was provided to D. L. Felder through U.S. Min- erals Management Service Cooperative Agreement 14-35-0001-30470, U.S. Fish and Wildlife Service Cooperative Agreement 14-16-0009-89-96— Task Order No. 6, NOAA Louisiana Sea Grant College Progi-am Grant RyCFB-21. and U.S. Department of Energy Grant No. DE-FG02-97ER12220. Literature cited Aiken, D. E., and S. L. Waddy. 1989. Allometric growth and onset of maturity in male American lobsters iHomarus americanus): the crusher propodite index. J. Shellfish Res. 8:7-11. Bauer, R. T. 1992. Testing generalization about latitudinal variation in reproduction and recruitment patterns with sicyoniid and caridean shrimp species. Invertebr Reprod. Dev. 22:193- 202. Claxton, W. T., and C. K. Govind. 1994. Chela function, morphometric maturity, and the mating embrace in male snow crab, Chionoecetes opilio. Can. J. Fish. Aquat. Sci. 51:1110-1118. Devine, C. E. 1966. Biology of Callianassa filholi Milne-Edwards 1878 (Crustacea, Thalassinidea). Trans. R. Soc. N.Z., Zool. 8:93-110. Dumbauld, B. R., D. A. Armstrong, and K. L. Feldman. 1996. Life-history characteristics of two sympatric thalas- sinidean shrimps, Neotrypaea californiensis and Upogebia pugettensis. with implications for oyster culture. J. Crust. Biol. 16:689-708. Felder, D. L. 1978. Osmotic and ionic regulation in several western At- lantic Callianassidae (Crustacea, Decapoda, Thalassin- idea). Biol. Bull. 1.54:409-429. 1979. Respiratory adaptations of the estuarine mud shrimp, Callianassa jamaicense (Schmitt, 1935) (Crustacea, Deca- poda, Thalassinidea I. Biol. Bull. 157:125-137. Felder. D. L., and R. B. Griffis. 1994. Dominant infaunal communities at risk in shoreline habitats: burrowing thalassinid Crustacea. Outer Conti- nental Shelf (OCS I Study MMS 94-0007. U.S. Dep. Inte- rior, Minerals Manage. Ser%'ice, Gulf of Mexico OCS Re- gional Office, New Orleans, LA, 87 p. Felder, D. L., and D. L. Lovett. 1989. Relative growth and sexual maturation in the estua- rine ghost shrimp Callianassa louisianensis Schmitt, 1935. J. Crust. Biol. 9:540-553. Felder, D. L., and R. B. Manning. 1997. Ghost shrimps of the genus Lepidophthalmus from the Caribbean region, with description of L. richardi. new species, from Belize (Decapoda: Thalassinidea: Callian- assidae). J. Crust. Biol. 17:309-331. Felder, D. L.. and S. de A. Rodrigues. 1993. Reexamination of the ghost shrimp Lepidophthalmus louisianensis (Schmitt, 1935) from the northern Gulf of Mexico and comparison to L. siriboia. new species, from Brazil (Decapoda: Thalassinidea: Callianassidae). J. Crust. Biol. 13:357-376. Felder, J. M., D. L. Felder, and S. C. Hand. 1986. Ontogeny of osmoregulation in the estuarine ghost shrimp Callianassa jamaicense var louisianensis Schmitt (Decapoda, Thalassinidea). J. E.xp. Mar Biol. Ecol. 99:91- 105. Forbes, A. T. 1973. An unusual abbreviated lan'al life in the estuarine prawn Callianassa kraussi (Crustacea: Decapoda: Thalas- sinidea). Mar Biol. 22:361-365. 1974. Osmotic and ionic regulation in Callianassa kraussi Stebbing (Crustacea, Decapoda, Thalassinidea). J. Exp. Mar Biol. Ecol. 16:301-311. 1977. Breeding and growth of the burrowing prawn Callia- nassa kraussi Stebbing (Crustacea. Decapoda. Thalas- sinideal. Zool. Afr 12:149-161. 1978. Maintenance of non-breeding populations of the es- tuarine prawn Callianassa kraussi (Crustacea. Anomura, Thalassinidea). Zool. Afr. 13:33-40. Grey, K. A. 1979. Estimates of the size of first maturity of the western 540 Fishery Bulletin 97(3), 1999 rock lobster, Panultrus cygnus, using secondary sexual characteristics. Aust. J. Mar. Freshwater Res. 30:785- 791. Gu, H., P. B. Mather, and M. F. Capra. 1994. The relative growth of chelipeds and abdomen and muscle production in male and female redclaw crayfish, Cherax quadricannatus von Martens. Aquaculture 123:249-257. Hailstone, T. S., and W. Stephenson. 1961. The biology of Callianassa (Trypaea) australieiisis Dana 1852 (Crustacea, Thalassinidea). Univ. Queensl, . Pap. Dep. Zool. 1(121:259-285. Hartnoll, R. G. 1974. Variation in growth pattern between some second- ary sexual characters in crabs (Decapoda BrachyuraK Crustaceana 27:131-136. 1978. The determination of relative growth in Crustacea. Crustaceana 34:281-293. 1982. Growth. In L. G. Abele (ed. I, The biology of Crusta- cea, vol. 2. Embryology, morphology, and genetics, p. 111- 1 16. Academic Press, New York, NY. Hintz, J. L. 1995. NCSS 6.0.1-3 user's manuals I-III. Number Cruncher Statistical Systems, Kaysviile, UT, 2204 p. Huxley, J. S. 1932. Problems of relative growth. Dial Press, New York, New \'ork. NY, 276 p. Lemaitre, R., and S. de A. Rodrigues. 1991. Lepidophthalmus siniiensis: A new species of ghost shrimp (Decapoda: Thalassinidea: Callianassidae) of im- portance to the commercial culture of penaeid shrimps on the Caribbean coast of Colombia, with observations on its ecology. Fish. Bull. 89:623-630. Lovett, D. L., and D. L. Felder. 1989. Application of regression techniques to studies of relative growth in crustaceans. J. Crust. Biol. 9:529-539. MacGinitie, G. E. 1934. The natural history of CaHianassa catiforniensis Dana. Am. Midi. Nat. 15:166-177. Manning, R. B. 1975. Two methods for collecting crustaceans in shallow water. Crustaceana 29:317-319. Manning, R. B., and D. L. Felder. 1990. Revision of the American Callianassidae ( Crustacea: Decapoda: Thalassinidea). Proc. Biol. Soc. Wash. 104:764- 792. McEdward, L. R., and F. Chia. 1991. Size and energy content of eggs from echinoderms with pelagic lecithotrophic development. J. Exp. Mar Biol. Ecol. 147:95-102. Mouton, E. C., Jr., and D. L. Felder. 1995. Reproduction of the fiddler crabs Uca longisignalis and Uca spinicarpa in a Gulf of Mexico salt marsh. Estuaries 18:468-481. Mukai, H., and I. Koike. 1984. Behavior and respiration of the burrowing shrimps Upogebia major (De Haan) and Callianassa japonica (De Haan). J. Crust. Biol. 4:191-200. Nates, S. F., and D. L. Felder. 1998. Impacts of burrowing ghost shrimp, genus Lepido- phthalmus, on penaeid shrimp culture (Crustacea: Decapoda: Thalassinidea). J. World Aqua. Soc. 29:189-211. Nates, S. F., D. L. Felder, and R. Lemaitre. 1997. Comparative larval development in two species of the burrowing ghost shrimp genus Lepidophthalmus ( (Crustacea: Decapoda: Callianassidae). J. Crust. Biol. 17:497-519. Pearse, A. S. 1945. Ecology of Upogebia affinis (Say). Ecology 26:303- 305. Phillips, P. J. 1971. Observations on the biology of mudshrimps of the genus Callianassa (Anomura: Thalassinidea) in Missis- sippi Sound. Gulf Res. Rep. 3:165-196. Pillai, G. 1990. Notes on the chelae of the mangrove-lobster Thalassma anomala (Decapoda, Thalassinidae). Crus- taceana 59:87-95. Pohl, M. E. 1946. Ecological observations on Callianassa major Say at Beaufort, North Carolina. Ecology 27:71-80. Poore, G. C. B., and T. H. Suchanek. 1988. Glypturus motupore, a new callianassid shrimp (Crus- tacea: Decapoda) from Papua New Guinea with notes on its ecology. Rec. Aust. Mus. 40:197-204. Posey, M. H. 1987. Effects of lowered salinity on activity of the ghost shrimp Callianassa californwnsis. Northwest Sci. 61:93-96. Robertson, W. D., and A. Kruger. 1994. Size at maturity, mating and spawning in the portunid crab Scylla serrata (Forskal) in Natal, South Af- rica. Estuanne Coastal Shelf Sci. 39:185-200. Rodrigues, S. de A. 1976. Sobre a reprodu^ao, embriologia e desenvolvimento larval de Callichirus major Say, 1818 (Crustacea, Deca- poda, Thalassinidea). Bolm. Zool.. Univ. S. Paulo 1:85- 104. Rowden, A. A., and M. B. Jones. 1993. Critical evaluation of sediment turnover estimates for Callianassidae (Decapoda: Thalassinidea). J. Exp. Mar Biol. Ecol. 173:265-272. 1994. A contribution to the biology of the burrowing mud shrimp, Callianassa subterranea (Decapoda: Thalas- sinidea). J. Mar Biol. Assoc. (U.K.) 74:623-635. 1995. The burrow structure of the mud shrimp Callianassa subterranea (Decapoda: Thalassinidea) from the North Sea. J. Nat. Hist. 29:1155-1165. Sastry, A. N. 1983. Ecological aspects of reproduction, //i J. F Vernberg and W. B. Vernberg (eds.). The biology of Crustacea, vol. 8, Environmental adaptations, p. 179-270. Academic Press, New York, NY'. Steele, D. H. 1988. Latitudinal variations in body size and species di- versity in marine decapod crustaceans of the continental shelf Int. Rev Gesamten. Hydrobiol. 73:235-246. Tamaki, A., and B. Ingole. 1993. Distribution of juvenile and adult ghost shrimps, Callianassa japonica Ortmann (Thalassinidea), on an in- tertidal sand flat: intraspecific facilitation as a possible pattern-generating factor .J. Crust. Biol. 13:175-183. Thompson, R. K., and A. W. Pritchard. 1969. Respiratory adaptations of two burrowing crusta- ceans, Callianassa californiensis and Upogebia pugettensis (Decapoda. Thalassinidea). Biol. Bull. 136:247-287. Tunberg, B. 1986. Studies on the population ecology of Upogebia deltaura ( Leach )( Crustacea, Thalassinidea). Estuarine Coastal Shelf Sci. 22:753-765. Vaugelas, J. de, B. Delesalle, and C. Monier. 1986. Aspects of the biology o{ Callichirus arniatus (A. Milne Edwards, 1870) (Decapoda, Thalassinidea) from French Polynesia. Crustaceana 50:204-216. Nates and Felder: Growth and maturation of Lepidophthalmus sinuensis 541 Witbaard, R., and G. C. A. Duineveld. truncata and their geochemical impact in the sea bed. 1989. Some aspects of the biology and ecology of the burrow- Nature 382:619-622. ing shrimp Callianasaa suhterranea I Montagu i (Thalas- Zimmerman, T. L., and D. L. Felder. sinidea) from the Southern Sea. Sarsia 74:209-219. 1991. Reproductive ecology of an intertidal brachyuran Ziebis, W., S. Forster, M. Huettel, and B. Jorgcnsen. crab, Sesarma sp. (nr. reticulatum). from the Gulf of 1996. Complex burrows of the mud shrimp Callianassa Mexico. Biol. Bull. 181:387-401. 542 Abstract.— The frequency with which dolphins from the northeastern offshore stock of the pantropical spotted dol- phin, Stenella attenuata, experience chase and capture by tuna purse-sein- ers in the eastern topical Pacific Ocean was estimated by comparing dolphin school-size frequencies in sighting data from research vessel observer records, with those recorded in set data by tuna vessel observers. The objective of the S study was to provide a preliminary ba- sis for estimating stock-wide effects of fishery-induced disturbance in these dolphins. Our analyses indicate two major characteristics for this stock: first, cap- ture frequency appears to increase rap- idly with increasing school size, and second, approximately half of the stock at any given time occurs in schools smaller than those apparently pre- ferred by purse-seiners. This implies that if individual dolphins have a pref- erence for associating with schools of a particular size, then individuals asso- ciating primarily with large schools would be subjected to chase and cap- ture much more frequently than those associating with small schools. How- ever, because the largest schools are relatively rare and account for a small percentage of individuals, the majority of dolphins in the stock would experi- ence relatively few captures per year, although some would experience a high rate. It is not known whether dolphins do indeed exhibit such a preference, or if instead individuals associate with schools from a wide range of sizes at different times. Schools of 1000 or more dolphins are estimated to be set on approximately once a week each on average, but such schools are estimated to represent just under one tenth of the animals in the northeastern offshore stock. Schools set on most often by tuna purse-seiners, containing from about 250 to 500 dol- phins, are estimated to be set on be- tween two and eight times each per year and are estimated to include ap- proximately one third of the stock. An estimated one half of the stock occurs in schools smaller than 250 animals; schools of this size are estimated to be set on less than twice per year each. Capture rate as a function of school size in pantropical spotted dolphins, Stenella attenuata, in the eastern tropical Pacific Ocean Peter C. Perkins Elizabeth F. Edwards Southwest Fisheries Science Center 8604 La Jolla Shores Dr La Jolla, California 92038 E-mail address (for P C Perkins) peter gcaliban ucsd edu Manuscript accepted 20 July 1998. Fish. Bull. 97:542-554 (1999). Tuna fishermen in the eastern tropi- cal Pacific Ocean (ETP) commonly catch large yellowfin tuna ( Thunnus albacares) by locating a school of dolphins visually and then sur- rounding it with a large purse-seine net in order to capture the tuna that are often closely associated with ETP dolphin schools. The dolphins are released from the net and the tuna are then loaded aboard (NRC, 1992). This method, known as "fish- ing on dolphin," historically has been a significant cause of dolphin mortality (NRC, 1992) but has also recently been suggested as a signifi- cant cause of fishery-related physi- ological stress in the dolphins in- volved, perhaps to the point of caus- ing unobserved mortality or changes in reproductive success (e.g. Myrick and Perkins, 1995). Although it has not been possible to measure physiological stress di- rectly in these dolphins, it is pos- sible to use existing data to estimate how often an animal experiences chase and capture. Capture fre- quency provides at least a rough measure of the amount of fishery- induced disturbance that dolphins affected by the ETP tuna fishery experience. Here, we estimate cap- ture frequency for the northeastern offshore stock of the pantropical spotted dolphin {Stenella attenuata ) (Dizon et al., 1992). This is the spe- cies most commonly associated with tuna and historically most often used in fishing on dolphin (greater than 70*7^ of dolphin sets annually for about the last 30 years (e.g. lATTCM. A simple calculation (see "Discus- sion" section) leads to a rough esti- mate for the mean number of times an individual dolphin is set on per year of (number of dolphins set on) H- (number of dolphins ) = 8 times per year. However, simply knowing the overall average rate of capture is not sufficient to evaluate the poten- tial adverse effects on individuals because the rate for different ani- mals may vary widely, depending on a number of interrelated factors in- cluding school size, geographic lo- cation, time of year, and the amount of tuna associated with a school. In this paper, we investigate the effects of school size. Specifically, we show that large dolphin schools (more than several hundred animals) are much more likely to be captured than are small schools (less than one hundred animals) because of a tendency for fishermen to concen- trate their effort on larger schools, which tend to carry more tuna, and to virtually ignore smaller ones. However, this result does not di- rectly give the capture rate for an lATTC. 1992. Annual report: 1990. In- ter-American Tropical Tuna Commission, La Jolla, CA, 261 p. Perkins and Edwards: Capture rate as a function of school size for Stenella attenuata 543 800 are not shown. (B) Tuna vessel sets. 1986-90; 19 observations >4000 not shown. Tlis) W,.f, l^\./I^S) -n is). /her e w Js'. w eff - the line transect effective strip half- width for schools of size s ; and = the size-averaged effective strip halfwidth (Appendix D in Burnham et al., 1980). 546 Fishery Bulletin 97(3), 1999 Table 2 Summary of tuna vessel data by fleet, year, and stratum. Coverage is defined as the percentage of fishing trips, involving sets on dolphins by vessels over 400 tons, that carried a scientific observer. Coverages and annual numbers of trips are taken from lATTC Annual Reports (e.g. lATTC, see Footnote 1 in main text). Observed sets per trip is defined as the mean number of observed sets on northeastern offshore spotted dolphin schools for trips that involved sets on dolphins. Middle and inshore strata are defined in Figure 1. Fleet Year Coverage Observed sets per trip 28 Jul-10 Dec {"study period") Observed sets per trip annual Trips Inshore Middle Pooled Trips Inshore Middle Pooled U.S. 1986 41.7 25 9.60 1.96 11.6 43 14.4 4.53 18.9 1987 91.5 61 17.3 5.25 22.6 119 21.9 6.89 28.8 1988 57.6 33 13.7 4.88 18.6 76 13.0 4.86 17.9 1989 99.1 50 14.9 3.66 18.6 115 15.3 3.77 19.1 1990 100.0 14 16.3 3.21 19.5 73 12.3 3.62 15.9 Total 76.9 183 14.9 4.14 19.0 426 16.2 4.88 21.1 Intl. 1986 25.3 29 12.3 2.90 15.2 68 14.9 2.84 17.7 1987 28.3 44 15.8 0.409 16.2 82 17.9 0.988 18.9 1988 35.8 54 8.70 3.52 12.2 111 10.3 3.87 14.2 1989 37.0 57 13.6 2.61 16.2 141 13.3 2.41 15.7 1990 42.0 66 8.35 6.92 15.3 147 12.3 4.77 17.1 Total 34.3 250 11.4 3.60 15.0 549 133 3.18 16.5 Both u\,Js) and «' ., depend upon the data trunca- tion distance (Burnham et al., 1980), denoted by w and equal to 5.5 km in this analysis. ;r*(s) was esti- mated from the observed school sizes. However, to estimate ;rts), we also needed to estimate «' .is), as described in "Estimation" section. Anecdotal reports consistently imply that tuna vessel captains do not search for or set on dolphin schools at random when fishing on dolphin in the ETP. Because larger dolphin schools are observed to carry more tuna, they are presumably sought out and set upon preferentially, and the set data would have a selection bias towards large schools. We modeled the schools associated with purse-seine sets (both observed and unobserved) as a biased sample, with replacement, of unknown size from the true popula- tion of schools. To characterize these schools, both the total number of sets, A'^^.^,,, and the effective probability density from which their sizes were drawn, p(s), needed to be estimated, pis) represents the superposition of the tuna fishermen's school-size selection preference upon nis). Sizes were recorded for all sets on trips car- rying observers, and therefore there was no additional observer selection bias (in relation top(s)). Assuming a random selection of trips, we treated the observed sets as an unbiased subsample of size n^^^^ and estimated pis) directly from the observed sizes. There was some concern with serial correlation between sets (see "In- dependence of observations" section). Because the number of dolphin schools is not con- stant over time, we interpreted A^jj.;,,,,,/,, as the time- averaged expected number of schools, and in particu- lar did not use a finite population estimator. Simi- larly, we interpreted A,,.,,, as the expected number of sets rather than making finite population estimates of the actual realized number of sets.'* Estimation U.S. vessels— study period The observed school-size distributions from both the research vessel sighting data and from the tuna vessel set data were roughly lognormal in shape (Fig. 2). We estimated ;r*(s) and p(s) using an adaptive kernel density estimator on the logs of the observed school sizes and then trans- formed the estimated density back to the original scale (Silverman, 1986). This variable bandwidth algorithm was chosen in order to make reasonable density estimates in the right tails where there were few data, while not oversmoothing near the modes. We chose the bandwidth scaling parameters as a trade-off between smoothness and fit to the data. We treated the observer estimates (mean estimates in the case of research vessel data) as exact measure- ■* Technical details and a discussion of our estimators for Af,^^^,, and A^ ,, can be found in Perkins. P. C. and E. F. Edwards, 1997, SWFSC Admin. Rep. LJ-97-03, Southwe.st Fisheries Sci- ence Center, La Jolla CA, 36 p. Perkins and Edwards: Capture rale as a function of scfiool size for Stenel/a attenuata 547 ments and did not attempt to correct for the possi- bility of size estimation biases in either data set (see "Discussion" section) or use a deconvolution kernel to account for estimation variance. Our estimates of wJs) were based on modelling the inherent selection bias in the research vessel sighting data. We used a bivariate hazard-rate de- tection function in a size-dependent line transect analysis of the perpendicular sighting distances and sizes of the observed schools (Drummer and McDonald, 1987; Palka, 1993). Perpendicular dis- tances were binned to reduce the effect of rounding in the data. School sizes were not binned because we did not use a parametric model for their distribution. We define the average capture frequency for a school of size s as (n ■hool: Setting the observed counts iig^i,^,^i^ and n^^^^ equal to their expectation gives 2Li •■ff '^schools ~ ^['hchonlsl ■ ^sets ~ '^['^setsi ~ t trips'^ sets i N.. ■hoob and where L A f ' trips the total distance searched by the re- search vessels; the total area within the stock bound- aries; and the fraction of tuna vessel trips that carried an observer. Using the relationship between /rts) and if'^is), and these moment equations for n^^^^^,^, and n,^,,,., we esti- mated the capture frequency for a school of size s as Note that the factor w ,,- cancels out and only w As) eft •' eft remains. Because there were so few schools smaller than 100 animals set on by tuna vessels, and so few schools larger than 1000 animals sighted from research ves- sels, we restricted our analysis to schools from 100 to 1000 animals, and computed estimates of capture frequency at intervals of 100 animals. _ Stratification and pooling There were many school size data from tuna vessel sets (n , =3454), however, sets ' there were far fewer research vessel sighting data = 499) and these had an uneven spatial distri- bution. We decided that research vessel sightings were too sparse to stratify our estimates of ;r*(s) and wJs) geographically. We did make stratified estimates ofp(s), but this had very little effect in absolute terms on the estimates of capture frequency, and so we present only pooled estimates for simplicity. Similarly, we did not stratify by year in any of our estimates. Set data, trip sampling fractions, sight- ing data, and search effort were combined to make a single estimate of the average N^^^^ and N^^i^^^g^ over all years. Extrapolation to the international fleet and to annual estimates Data from individual sets came only from U.S. tuna vessels, so that estimating capture fre- quency due to the entire fleet required extrapolation. We made the assumption that a captain's preference for dolphin school sizes upon which to make sets did not vary with the vessel's country of origin and thus extrapolated our estimate of p(s) to the entire fleet. Our estimates of total numbers of target sets were based on separate observed counts and sampling frac- tions for the U.S. and the international fleets, sets sets sets "sets /■It/Si f trips Jlntl) Itr ilnth ips This expression was substituted in for '",(,(j //^rip.^i in our estimates of capture frequency for the com- bined fleet. For non-U. S. vessels, we had only the total number of target sets observed, and we were unable to estimate the variance in N'^""\ Thus, our sets _ estimates for the combined fleet do not include esti- mates of precision. Our estimates of capture frequency are strictly valid only for the study period. However, if we as- sume that the same patterns in school sizes and captain's preference for school sizes hold for the en- tire year, then the annual capture frequency can be estimated by using the corresponding annual set counts for U.S. and non-U. S. vessels. Because we had only the total number of target sets observed during times other than the study period, our estimates of annual capture frequency do not include estimates of standard error. Independence of observations Because of the geographically correlated nature of consecutive research vessel sightings or tuna vessel sets, successive school size observations from a single vessel may not have been independent. This is par- ticularly a concern for the set data, because of the possibility of repeated sets on the same school (see 548 Fishery Bulletin 97(3), 1999 "Discussion" section). Although dependence does not add a bias to our estimates, it does de- crease the effective sample size, which affects our estimates of precision. We accounted for this prob- lem by using bootstrap standard error estimates, and by defining our bootstrap resampling units so as to make them as independent as possible while keeping a reasonably large sample size. For research vessel data, we took days as the resampling unit; for tuna vessels, we resampled by trips. For each bootstrap iteration, we resampled from the research vessel data to achieve approximately the same amount of search effort in each stratum as was actually achieved. We resampled fi-om the tuna vessel data to achieve exactly the actual observed number of trips. Results Estimated distribution of school sizes Figure 2 shows the kernel estimates of the den- sities ;r*(s) andp(s). Both estimated densities were much smoother at large school sizes than at small school sizes. This is partially due to the variable bandwidth in the kernel estima- tor, but primarily due to the data themselves. Estimated effective strip halfwidth Figure 3 shows the estimated values for the ef- fective strip halfwidth as a function of school size. Because ;r*(s)o^ii'p/^s);r(s).uv/f's) represents the relative amount of "thinning" for schools of different sizes, i.e. Weff(s)lw is the probability of a school of size s being detected from the re- search vessel, given that it is within the trun- cation distance w. The estimated values indi- cate that approximately one third of schools of size 100 within the truncation distance (5.5 km) were missed by the research vessel observers, and essen- tially all schools of size 1000 were detected. The re- sult shown in Figure 3 is, qualitatively at least, par- tially constrained by the bivariate line transect model, i.e. if the data indicate dependence of detect- ability upon school size, then the parametric form for tiV/f s] = [t7iit)dt [tK(t)dt //(s) = (s, where I\s, > s] ■ [1, if.s, >s [0, otherwise and the sums are over research vessel sightings. 550 Fishery Bulletin 97(3), 1999 Combining H (s) with the combined-fleet cap- ture frequency estimates (Fig. 5, upper cui-ve), schools of 1000 animals or greater were esti- mated to be set on at least once every ten days and contained an estimated 9% of dolphins (Fig. 6). Schools set on most often by tuna purse-sein- ers, containing from about 250 to 500 dolphins, were estimated to be set on between 2 and 8 times each per year on average; these schools repre- sented just under an estimated one third of the stock. An estimated one half of NE offshore spot- ted dolphins occurred in schools smaller than 250 animals; schools of this size were estimated to be set on less than twice per year each. We note that//(s) should not be used to quan- tify the school size preferences of individual dolphins. For example, although we estimated that schools larger than 1000 animals contained an estimated 9% of dolphins at any given time, this does not imply that the same 9% of dol- phins always made up such schools. Capture frequency for very large schools Although no kernel estimate of nis) was pos- sible for s greater than 1000 animals, the esti- mated detection probability for those schools was essentially one out to the truncation dis- tance w, making a rough calculation for capture frequency possible. Because of the rounding tendency of tuna vessel observers, we made an estimate for schools greater than or equal to 1000. With the assumption that the effective strip halfwidth is equal to the truncation dis- tance, an estimate of the average capture fre- quency due to the entire fleet is J^(large) N , {S> 1000) = ^Z^ N. lUS. large) schtmh ft US I y I trips I Inns bill 1 ■ips 2Li ^"■schools o US and Int'l fleet • 8 • CD 0) >■ Q. • c o / times set 20 • o / / E 3 / US fleet z • o . / ^^ • ^^* ^ • ""^^^ ^-^ • o %^=^' — -' 100 200 300 400 500 600 700 800 900 1,000 School size Figure 5 Estimated annua! capture frequency as a function of dolphin school size for schools of northeastern offshore spotted dolphins. The estimates are of the average number of times a school was set on each year by tuna vessels in the ETP purse-seine fleet for the years 1986-90. The lower curve shows the number of sets due to U.S. vessels only; the upper curve shows the number of sets due to U.S. and non-U. S. vessels combined. Estimates of stan- dard error were not possible for these estimates. The estimated average capture frequency for these very large schools was 51.3 sets per year, or just un- der once a week. Discussion Capture rates for individual dolphins To interpret these capture frequency results in terms of individual dolphins, we must consider the size range of the schools with which a given individual tends to associate. If one assumes that dolphins have a strong fidelity for a characteristic school size, then the above results indicate that a fixed but relatively small percentage of the dolphin population was con- sistently subjected to a high rate of capture in purse- seine nets, whereas the majority of dolphins were subject to relatively little disturbance from the fish- ery However, little is known about the spatial and temporal dynamics of dolphin schools and their sizes, and a range of other assumptions are possible. If school membership is completely fluid and dol- phins mix perfectly among schools, then over the long term, all dolphins would experience the same cap- ture rate. We made a rough estimate of this rate by estimating the total annual number of dolphins set on and the total number of dolphins. Using data from Table 2, we made a rough estimate of 7610 for the Perkins and Edwards: Capture rate as a function of schiool size for Stenella attenuala 551 mean annual number of sets on NE offshore spotted dolphins during the period of this study. From tuna vessel observer data, an estimate for the mean school size for those sets is 773 animals. When combined with an estimate for the total number of NE offshore spotted dol- phins (Wade and Gerrodette, 1993), this gives (7610 X 773 dolphins set on) ^ (731,000 dolphins) = 8.04 sets per dolphin per year. The true picture certainly lies between these two extremes. On the other hand, if the compo- sition and spatial location of some large schools are static over periods of weeks or longer, then animals in those schools could be subject to short-term capture rates even higher than those of our estimates because of the clustered distri- bution of fishing effort. Geographic stratification Holt et al. ( 1987) partitioned the ETP into sev- eral geographic strata primarily on the basis of spotted dolphin density as observed from tuna vessels during years prior to the study period. In an alternate analysis for capture frequency, we used Holt et al.'s partition to fit separate densities for p(s) in each of two strata (Fig. 1), and used separate counts n , and n , , . The ^ ^'t=^s schools estimates of p(s) from the two strata were sig- nificantly different (Kolmogorov-Smirnov good- ness-of-fit test, P=0.002), but primarily at smaller school sizes, less than 200 animals, and this stratification made little difference in ab- solute terms from the unstratified estimates of capture frequency. The similarity in capture frequency estimates between strata indicates that fishing pressure was approximately pro- portional to dolphin school density. We did not stratify geographically to estimate ;r*(s ) or u\,y{s) because we found that the number of ob- servations in the middle stratum (n , ,=81) was too small to allow stratification and still have reason- able precision. A Kolmogorov-Smirnov test and Q-Q plots indicated that there was no substantial differ- ence (P=0.62) in ;r*(s) between strata for the research vessel observers. On the other hand, we fitted the bivariate line transect model to data from the two strata separately and found that the estimate of i<-\,fyis) for the middle stratum was 10-20^^^ smaller than that for the inshore stratum, depending on school size. However, there were few data on which to base either result. One reason why u\,f-As) might actually have differed between the two strata was a difference in observed sea state conditions; a higher average Beaufort sea state was reported in the CO d d d CM d o d 100 = 300 .J = 500 i =800 i = 1 ,000 10 20 30 Minimum number of times set on per year 40 Figure 6 Estimated percentage of northeastern offshore spotted dolphins subject to different levels of capture frequency. The horizontal axis represents the minimum number of times a dolphin school is set upon per year by U.S. and non-U. S. tuna vessels in the ETP purse-seine fleet, for the years 1986-90. The vertical axis represents the estimated fraction of the stock (not the fraction of schools I subject to at least that rate of being set upon, s is the minimum school size accounting for that percentage. middle stratum. The practical impact is that our es- timates of capture frequency may have been overin- fiuenced by data from the inshore stratum. Observer size-estimation errors Our statistical model for the school size data included terms for selection biases, that is, which schools were included in the sighting or set data. However, there was also a potential for observer size-estimation bi- ases. That is, given a sighting of, or a set on, a spe- cific school, an observer had to estimate size of the school. The results presented here treated the ob- server estimates as exact counts. We did not include an error term for size estimates in either the kernel density estimates of ;r*(s) and p(s) or the bivariate line transect estimates of u' ,,(s). 552 Fishery Bulletin 97(3), 1999 An observer size estimation bias would scale or otherwise deform those estimates of if'' is) or pis) (or both), depending on whether the bias was propor- tional to size or was more complex. Even if the ob- servers were unbiased in their individual estimates, estimation variance would still increase both tails in the density estimates. Thus, if research vessel observers and tuna vessel observers consistently made different errors in estimating school sizes, then S the trend in our estimates of capture frequency, e.g. Figure 4, could have been in part or entirely due to those errors. Gerrodette and Perrin^ studied dolphin school size- estimation errors for research vessel observers by ground-truthing observer estimates against aerial photo counts of the same school. They found that the counts from a single observer could be modeled as lognormally distributed given the true school size. They also found that a given observer in the study might have a substantial positive or negative bias. Using their photo and observer dataset, we fitted a lognormal model for the geometric means of the observer estimates from each sighting.'' The fit indi- cated that the observers had essentially no bias at a true school size near 100, but that there was a nega- tive bias of 21% at a school size of 1000. Because the lognormal is a skewed distribution, this mean bias cor- responds to essentially no median bias. The estimated CV for the estimates, given the true size, was 48%. Given this information, it would have been pos- sible in theory to correct the research vessel observer estimates for bias. However, we did not make that correction because the corresponding correction for the tuna vessel set data was not possible in the ab- sence of a suitable ground-truth study. Significant differences between observers and observing condi- tions on research vessels and tuna vessels precluded the assumption that any size-estimation biases are similar in the two types of data. Spatial distribution of schools and school sizes Our analysis can be taken to imply full spatial mix- ing, that is, all schools of a given size within the stock boundaries (or within each stratum for the strati- fied case) have the same probability of being set upon. A more realistic model is that some schools have a higher or lower probability depending not only on their size but also on their geographic location in relation to areas of high school density or high fish- ing effort. Other factors, such as seasonal effects and the amount of associated tuna, are also interrelated with location in determining the rate of capture for a given school. Because we included only limited spa- tial information in our model, the appropriate inter- pretation of our results is that we estimated an av- erage probability of being set upon, as a function of school size, for schools within the stock boundaries (or each stratum). Observer experience suggests that pressure from fishing on dolphin can reduce average dolphin school size, i.e. areas of high fishing effort tend to have smaller schools. ' This decrease in school size may be a result of chase and capture operations during sets intentionally or unintentionally splitting schools into smaller subgroups.*^ However, we did not find any indication of such a trend in our NE offshore spotted dolphin school-size data. We concluded that if fish- ing pressure did affect spotted dolphin school size, its effects may have been masked by size selection in the tuna vessel set data and by the relatively lim- ited number of observations in the research vessel sighting data. Dolphin schools Variation in school size over time The simplest in- '' terpretation of this analysis would assume that a dolphin school is a fixed entity that does not change in size. If, on the other hand, a school is not a well- defined entity over any length of time, i.e. schools often fragment and reaggregate (e.g., Scott and Cattanach, 1998), then defining capture frequency for anything other than an individual dolphin be- comes problematic. The superpopulation model that we used is one way to account for this fiuid nature of dolphin schools. In particular, the research and tuna vessel school-size data represent time-averaged samples, i.e., averages over repeated realizations fi-om the superpopulation. Estimated capture frequencies can be interpreted in terms of short-term rates. Species composition Most schools in both the re- search vessel and tuna vessel data included not only NE offshore spotted dolphins, but other species as well, primarily spinner dolphins iStenella longir- ostris). We did not differentiate between pure and 5 Gerrodette, T. and C.Pcrrin. 1991, SWFSC Admin. Rep. LJ- 91-36, Southwest Fisheries Science Center, La Jolla CA, 74 p. * Details of this exploratory analysis can be found in Perkins, P. C. and E. F. Edwards, 1997, SWFSC Admin. Rep. LJ-97-03, Southwest Fisheries Science Center, La Jolla CA, 36 p ' Rasmussen, R. 1997. Southwest Fisheries Science Center, La Jolla, CA. Personal commun. ** Hall, M. 1997. Inter-Amencan Tropical Tuna Commission, La Jolla, CA. Some limited data have been collected to study .school fragmentation and reaggregation (Perrin et al., 1979; Scott, M. 1997. Inter-American Tropical Tuna Commission, La Jolla, CA. Personal commun. Perkins and Edwards: Capture rate as a function of scfiool size for Stenella attenuata 553 mixed schools in our analysis. Thus, we took as our population of schools not just those composed purely of NE offshore spotted dolphins, but all schools con- taining them. School sizes were taken as the total number of animals in each school. This approach would not have been appropriate if we had been es- timating a stock-specific abundance (e.g. Wade and Gerrodette, 1993). However, as long as there is no bias in the species composition of schools that are set on, our approach is valid. An exploratory data analysis indicated that the distribution of species proportions was very similar for both research and tuna vessel data. There was some indication that pure spotted schools tended to be smaller on average than mixed spotted-spinner schools. We did not pursue this be- cause it did not affect our results. Encounter rate for very large schools Inspection of Figure 2 raises the question of why so few very large schools { 1000 animals or gi'eater) were sighted from the research vessels when so many were set upon by tuna vessels. Only five schools ( f/r of sightings) in that range were reported by research vessel obsei-vers, and the largest was estimated to be 2617 animals. In that range, 896 schools (26"^ of observed sets) were reported set upon by tuna vessel observers, and 97 were estimated to be larger than 2617 animals. These largest schools from the set data did tend to include slightly higher percentages of species other than spotted dolphins. However, they were still primarily made up of spotted dolphins (just over an estimated 7Q'7( on average), and it was not the case that they were due to an association with large groups of, for example, common dolphins (Del- phinus delphis). which are known to form very large schools (e.g. Edwards and Perrin, 1993). At least four explanations for this apparent dis- crepancy are possible. First, this may simply reflect the tuna vessel captains" preference for setting on large schools. Second, the difference may be due to relative bias in size estimation between the two types of observers, as discussed earlier. However, to explain all of the difference, the two sets of obsei'vers would have to differ on average by a factor of five in their estimates. Third, the research vessels may have missed a relatively rare segment of the population of schools, which the tuna vessels are able to seek out with a much greater search effort and a nonran- dom search strategy. Fourth, some of these large ob- servations in the set data may have been from in- tentionally repeated sets on the same schools. There is evidence in the tuna vessel obsei^ver data for both of these last two explanations, i.e., that localized ar- eas of high density or school size (or both) may exist and that repeated sets on a single school may occur. Conclusions The results of this study indicate that tuna purse- seiners in the ETP fishing on NE offshore spotted dolphins have a strong preference for setting on larger than average dolphin schools, and that such schools were subject to being set on at a much higher rate than were smaller schools. Specifically, the larg- est schools considered, those of 1000 or more ani- mals, were estimated to be set on approximately once every week, whereas the smallest schools considered, those of 100 animals, were estimated set on less than once a year. Our estimated capture rates should be taken as averages for a given school size and do not account for variation due to other factors, such as geographic location or the amount of associated tuna. Also, although we estimated rates in terms of sets per year, we do not assert that the short-term cap- ture rate for a given school is constant, i.e. that sets occur at evenly spaced intervals throughout the year. For example, relatively few sets are made "on dol- phin" in the NE offshore stock range during June and July (e.g. Edwards and Perkins, 1998). These results do not account for any errors in esti- mation of dolphin school size. Although potential er- rors in school-size estimates made by research ves- sel observers were investigated, no corresponding study of potential errors for tuna vessel observer es- timates was possible. To draw conclusions about capture frequency for an individual dolphin, we must consider the size range of the schools with which a given individual tends to associate. Our results imply that dolphins associating primarily with large schools will be sub- jected to capture much more often than individuals associating primarily with small schools. However, we also estimated that the largest schools are rela- tively rare and account for a minority of the total number of individual dolphins at any given time. These results may imply that a fixed but relatively small percentage of the dolphin population was con- sistently subjected to a high rate of capture in purse- seine nets but that a majority of dolphins occur in schools smaller than those apparently preferred by purse-seiners, and experience relatively few captures per year. However, little is known about the spatial and tem- poral dynamics of dolphin schools and their sizes, and other conclusions are possible. If dolphins asso- ciate with a wide range of school sizes, then the cap- ture rates for individual dolphins would tend to "av- 554 Fishery Bulletin 97(3), 1999 erage out" and thus would vary less than the range of capture rates for schools. On the other hand, dif- ferences between schools in factors other than size could lead to short-term individual capture rates that are even higher than our estimates. Acknowledgments ^ We thank M, Hall, M. Garcia, C. Lennert, and M. Scott of the Inter- American Tropical Tuna Commis- sion, and W. Armstrong, J. Barlow, R. Holt, A. Jack- son, R. Rasmussen, and K. Wallace of the SWFSC, for the benefit of their field experience and their tech- nical knowledge of the tuna vessel obser\'er program, the NMFS research cruises, and the related data- bases. Special thanks to T. Gerrodette of the SWFSC and D. Palka of the NEFSC for theoretical and prac- tical advice on the bivariate hazard rate model. We also thank J. Barlow, B. Curry, T Gerrodette, M. Hall, P. Kleiber, and two anonymous reviewers for their comments on draft versions. Finally, we gratefully acknowledge all of the scientists, observers, NOAA officers and crew, and fishermen who participated in the collection of the research vessel observer data and the tuna vessel observer data. Literature cited Burnham, K. P., D. R. Anderson, and J. L. Laake. 1980. Estimation of density from line transect sampling of biological populations. Wildlife Monograph 72, supple- ment to Journal of Wildlife Management 44. 202 p. Dizon, A. E.. W. F. Perrin, and P. A. Akin. 1992. Stocks of dolphins iStenella spp. and Delphinus delphis'i in the eastern tropical Pacific: a phylogeographic classification. U.S. Dep. Commer.. N'OAA Tech. Rep. NMFS 119, 20 p. Drummer, T. D., and L. L. McDonald. 1987. Size bias in line transect sampling. Biometrics 43:13-21. Edwards, E. F., and P. C. Perkins. 1998. Estimated tuna discard from dolphin, school, and log sets in the eastern tropical Pacific Ocean. 1989-92. Fish Bull. 96:210-222. Edwards, E. F., and C. Perrin. 1993. Effects of dolphin group type, percent coverage, and fleet size on estimates of annual dolphin mortality derived from 1987 U.S. tuna-vessel observer data. Fish. Bull. 91:628-640. Hill, P. S., R. C. Rasmussen, and T. Gerrodette. 1991. Report of a marine mammal survey of the eastern tropical Pacific aboard the research vessel David Starr Jordan, July 28-December 6, 1990. U.S. Dep. Commer., NOAA Tech. Memo. NMFS NOAA-TM-NMFS-S\\TSC-158, Southwest Fisheries Science Center. La Jolla, CA, 133 p. Holt, R. S., T. Gerrodette, and J. B. Cologne. 1987. Research vessel survey design for monitoring dolphin abundance in the eastern tropical Pacific. Fish. Bull. 85:43.5-446. lATTC (Inter-American Tropical Tuna Commission). 1989. Incidental mortality of dolphins m the eastern tropi- cal Pacific tuna fishery, 1979-1988: a decade of the Inter- American Tropical Tuna Commission's scientific technician program. Working document 2 from the tuna-dolphin work- shop held 14-16 March, 1989 in San Jose, Costa Rica. Inter-.-Vmencan Tropical Tuna Commission, La Jolla. CA, 90 p. 1991. Tuna-dolphin program field manual. Inter-Ameri- can Tropical Tuna Commission, La Jolla. CA, 170 p. Jackson, A. R. 1993. Summary of 1989 U.S. tuna-dolphin observer data. U.S. Dep. Commer, NOAA Tech. Memo. NMFS NOAA-TM-NMFS-SWFSC-183, Southwest Fisheries Sci- ence Center. La Jolla. CA. 31 p. Mangels, K. F. and T. Gerrodette. 1994. Report of cetacean sightings during a marine mam.- mal survey of the eastern topical Pacific on the research vessels David Starr Jordan and MacArthur. July 28-No- vember 2, 1992. U.S. Dep. Commer, NOAA Tech. Memo. NMFS NO.A\-TM-NMFS-SWFSC-200, Southwest Fisher- ies .Science Center. La -Jolla. CA, 74 p. Myrick, A.C., Jr., and P. C. Perkins. 1995. Adrenocortical color darkness and correlates as in- dicators of continuous acute premortem stress in chased and purse-seine captured male dolphins. Pathophysiology 2:191-204. NMFS (National Marine Fisheries Service). 1992. Purse seine obser\'er field manual. National Marine Fisheries .Ser\ice. Long Beach. CA. 258 p. NCR (National Research Council). 1992. Dolphins and the tuna industry. National Academy Press, Washington DC. 176 p. Orbach, M. K. 1977. Hunters, seamen, and entrepreneurs: the tuna seinermen of San Diego. Uni\'. California Press, Berke- ley CA, 304 p. Palka, D. L. 1993. Estimating density of animals when assumptions of line-transect surveys are violated. Unpublished Ph.D. diss.. Univ. California, San Diego, CA, 169 p. Perrin, W. F., W. E. Evans, and D. B. Holts. 1979. Movements of pelagic dolphins iStenella spp.) in the eastern tropical Pacific as indicated by results of tagging, with summary of tagging operations, 1969-76. U.S. Dep. Commer.. NO.AA Tech. Rep NMFS SSRF-737, 14 p. Scott, M. D., and K. L. Cattanach. 1998. Diel patterns in aggregations of pelagic dolphins and tunas in the eastern Pacific. Mar. Mamm. Sci. 14:401-428. Silverman, B. W. 1986. Density estimation for statistics and data analysis. Chapman and Hall, London, 175 p. Wade, P. R., and T. Gerrodette. 1993. Estimates of cetacean abundance and distribution in the eastern tropical Pacific. Rep. Int. Whal. Comm. 43:477-493. 555 Abstract.— Temporal and spatial vari- ability in growth and mortality rates of bay anchovy, Anchoa mitchilli. lar- vae was analyzed in Chesapeake Bay. Larvae were collected in cruises dur- ing June and July 1993. on transects spaced at 18.5-km (10 nmi) intervals over the entire bay. Growth and mor- tality rates were estimated in lower, mid, and upper bay regions and ana- lyzed in relation to environmental vari- ables, predators (biovolumes of the scvphomedusa Chn'saora quinquecirrha and the ctenophore Mnemiopsis leidyi). and larval prey (zooplankton abun- dances). Otolith increment analysis in- dicated that the mean baywide growth rate of larvae increased significantly from 0.59 mm/d in June to 0.72 mm/d in July. The baywide mortality rate of larvae declined from 0.41 (33.6'^f/d) in June to 0.23 (20.5'^f/d) in July In each m.onth. regional mortality rates were highest in the lower bay. Regionally, mortality ranged from a low of 0.14 (13.1'^?-/d I in the upper bay in July to a high of 0.54 (41.7%^) in the lower bay in June. Mortality rates declined with increasing larval size. Stage-specific survival was both size-specific and growth-rate dependent as indicated by trends in mortality (M), weight-specific growth iGl, and the M/G ratio. Growth rates were positively correlated with temperature and zooplankton abun- dance. Larval abundances, but not mor- tality rates, were negatively correlated with gelatinous predator biovolumes. Recruitment potential of bay anchovy was judged to be highest in July in the lower third of Chesapeake Bay. Al- though lower, production of anchovy prerecruits in June and in other Bay regions was substantial and contrib- uted significantly to prerecruit abun- dances in 1993. Regional and temporal variability in growth and mortality of bay anchovy, Anchoa mitchilli, larvae in Chesapeake Bay Gene C. Rilling Edward D. Houde University of Maryland Center for Environmental Science Chesapeake Biological Laboratory PO Box 38 Solomons, Maryland 20688-0038 Present address (for G C Rilling); Connecticut Department of Environmental Protection Office of Long Island Sound Programs 79 Elm Street, Hartford, Connecticut 06106-5127, E-mail address (for G C Rilling) chris rilling apostate ct us Manuscript accepted 13 July 1998. Fish. Bull. 97:555-569 ( 1999). In highly fecund fishes that spawn serially over a protracted season and a broad geographic range, vari- able patterns of cohort successes and failures may result that not only lead to fluctuating recruit- ments but that are difficult to de- tect in the absence of sampling pro- grams that are temporally and spa- tially intensive. We estimated tem- poral and regional variability in lar- val-stage growth and mortality of bay anchovy {Anchoa mitchilli). an abundant (Hildebrand and Schroe- der. 1928) and highly productive species (Newberger and Houde, 1995; Wang and Houde, 1995) in Chesapeake Bay. In exploratory analyses and simulations, Houde (1996, 1997b) demonstrated that variability in recruitment success of bay anchovy can occur when stage- specific mortality rates vary during early life. Recently, individual- based models of bay anchovy dy- namics in Chesapeake Bay have demonstrated how stage-specific mortality and growth processes may operate and how density-dependent regulation could dampen fluctua- tions in abundance iWang et al., 1997). Bay anchovy are an important component of the Chesapeake Bay food web. They are not commer- cially exploited but are a major prey of harvested species such as blue- fish (Pomatomus saltatrix), weak- fish (Cynoscion regalis), and striped bass (Morone saxatilis) (Hartman and Brandt, 1995) and may repre- sent up to 90*^ of piscivorous fish diets seasonally (Baird and Ulano- wicz, 1989). Spawning by bay an- chovy is widespread in the Bay and occurs over a broad range of tem- peratures and salinities (Dovel, 1971; Houde and Zastrow, 1991). Bay anchovy is a pelagic, serial spawner (Luo and Musick, 1991; Zastrow et al., 1991) that spawns most intensively from May through August in Chesapeake Bay, where it may account for 96-99^)?^ of fish egg and 67-88T of larval catches (Olney, 1983). During peak spawn- ing, densities of eggs frequently range from 10 to >1000/m'^ and den- sities of larvae from 1 to >100/m'^ (Olney, 1983; Dalton, 1987; Dorsey et al., 1996; MacGregor and Houde, 1996; Rilling and Houde, manu- script in review). Within a spawn- ing season, regions of highest egg and larval abundances may shift, as they did from the upper to the lower bay between June and July 1993 (Rilling and Houde, manuscript in review). Such shifts may have im- portant repercussions for bay an- chovy production and for production of other fish species and inverte- 556 Fishery Bulletin 97(3), 1999 N 39° 30' 39° 00' 38° 30' 38° 00' 37° 30' brate predators that rely upon bay anchovy as prey. Because bay anchovy eggs and lar- vae, and its gelatinous predators, peak in abundance during summer, temporally and spatially variable pre- dation is potentially a significant fac- tor controlling bay anchovy survival and recruitment. Results of meso- cosm experiments ( Cowan and Houde, 1993 ) have indicated that up to 20-40% of bay anchovy eggs and larvae in Chesapeake Bay during the peak spawning season may be consumed daily by jellyfish. Purcell et al. ( 1994) analyzed jellyfish gut contents and estimated that these predators could account for up to 21'7( of the daily egg mortality and 41'7f of the larval mor- tality of bay anchovy in Chesapeake Bay. In site-specific studies, Dorsey et al. (1996) estimated that jellyfish accounted for O-SS'/r/d of egg mortal- ity, and from 0 to IS'/f/d of yolksac larval mortality. Minor variability in daily mortal- ity or growth rates in a 30-50 d pe- riod during early life, can generate tenfold or greater differences in re- cruitment potential of marine fish (Gushing, 1975; Houde, 1987, 1989b). Previous studies in Chesapeake Bay have documented variability in an- chovy egg and larval abundances and yolksac larval dynamics at regional scales (Dorsey et al., 1996; MacGregor and Houde, 1996), but this study is the first to provide a comprehensive overview of larval dynamics for the entire Chesa- peake Bay. At the outset, we proposed that there are regional and temporal differences or patterns in growth and mortality rates of bay anchovy larvae in Chesapeake Bay. To test this hypothesis, we analyzed otolith microstructure to estimate and compare month-specific (June vs. July), and region-specific (three regions) growth and mortality rates. Incre- ments are deposited daily on sagittal otoliths of bay anchovy, providing a record of age (Fives et al., 1986; Leak and Houde, 1987; Castro and Cowen, 1991; Zastrow et al., 1991). Deposition of otolith increments begins on the third day after hatching and has been documented in laboratory experiments (Leak and Houde, 1987). Here, we discuss how regional or tem- poral variability in growth and mortality might in- fluence potential for recruitment and production. We 37° 00' Upper Mid Lower 77° 30' 77° 00' 76° 30' 76° 00' 75° 30' W Figure 1 Chesapeake Bay. Regions, transects, and station locations for ichthyoplankton samples, zooplankton samples, and CTD casts, 19- 22 June and 23-30 July 1993. compared growth and mortality rates of bay anchovy larvae in relation to abundances of gelatinous preda- tors and zooplankton (larval prey), and in relation to salinity and temperature. Overall, our objective was to determine which region(s ) of the Bay and what part of the spawning season contributed most to sur- vival and subsequent production of bay anchovy early-life stages. Materials and methods Ichthyoplankton was collected throughout Chesa- peake Bay during two baywide cruises, 19-22 June and 23-30 July 1993 (Fig. 1). Forty-six stations were sampled in June and 48 in July. Stations were on 15 transects spaced at 18.5-km ( 10 nmi) intervals from Rilling and Houde: Variability in growth and mortality of Anchoa mitchilli 557 the head of the Bay (39°25'N) to near the Bay mouth (ST^OS'N). Data were analyzed and compared in three regions: upper bay— transects 1-5 (39°25'N-38°45'N); mid bay— transects 6-10 (38°45'N-37°55'N); and lower bay— transects 11-15 (37°55'N-37°05'N). At each station, ichthyoplankton was collected in one net of an opening-closing, 60-cm bongo sampler with 280-(im meshes and preserved in ethanol. Two tows, of 2-min duration, were made at each station. The first tow was from within 1 m of bottom to the pycnocline, and the second was made from the pycnocline (or middepth when no pycnocline was present) to the surface. Sampling protocols and meth- ods to estimate densities and abundances of organ- isms are detailed by Rilling and Houde, manuscript in review); only brief descriptions are given here. Immediately before each tow, a conductivity-tem- perature-depth (CTD) cast was made from within 1.0 m of bottom to within 1.0 m of the surface to pro- vide depth profiles of temperature, salinity, and dis- solved oxygen. To make results comparable to those of Dorsey et al. (1996) and MacGregor and Houde (1996), temperature, salinity, and dissolved oxygen were examined at 3-m depth. Zooplankton from ei- ther three or four designated depths was sampled in 10-L Niskin bottles and collected on 35-/;m mesh. Gelatinous zooplankters from each ichthyoplankton tow were counted and their biovolumes recorded. In the laboratory, anchovy lar\'ae were measured to the nearest 0.1 mm standard length (SL). Lengths were corrected for shrinkage during collection and preservation (Theilacker, 1980; Leak, 1986). Small larvae of bay anchovy may be extruded through net meshes (Leak and Houde. 1987). Therefore, we ap- plied a regression method to adjust abundances of <5.5 mm SL larvae collected in the 280-|im net meshes. The regression, which adjusted abundances of larvae upward by factors of 2.3 (at 2.0 mm), 1.9 (at 3.0 mm), 1.6 (at 4.0 mm), and 1.2 (at 5.0 mm lar- vae), was derived from comparisons of length-specific abundances in paired tows of 53-)im and 280-|.im mesh bongo nets made in Chesapeake Bay under condi- tions similar to those during this survey (MacGregor, 1994; Rilling and Houde, manuscript in review). Otolith analysis Otolith microstructure was analyzed to estimate age, growth, and mortality. In the present study, sagittal otoliths from 509 larvae were examined. Otoliths from representative samples of larvae from each re- gion of the Bay were examined for each cruise. Each larva in the otolith analysis was measured to the nearest 0.1 mm SL. Otoliths from larvae of 2.0 to 25.0 mm SL were mounted in "Epon" under a cover slip, and heated for 24 h at 60°C to harden the epoxy (Secor et al., 1991 ). Otolith increments were counted on two separate occasions under a compound light microscope at 600 to lOOOx magnification by one reader (Rilling). The mean of the two increment counts plus two days was the estimated age. Growth rates Growth in length of larvae (mm/d) was estimated from the slopes of the linear regressions of shrink- age-adjusted lengths (SL) on ages from daily otolith- increment analysis: L, = a + gt. where L, = standard length (mm) at age t (d); t = age (d) = otolith increment count plus two days; g = growth rate (mm/d); and a = y-intercept, the estimated length (mm) SL at hatch. Gompertz growth models (Bolz and Burns, 1996) also were fitted to the data for each region and cruise. The fits were no better than those for the linear model and were not considered further in our analysis. Larval lengths were converted to dry weights (g) from a weight-length relationship: W=0.1550xL3 5307_ where W = dry weight (g); and L = mm SL. Rates of growth in weight then were estimated from an exponential model, fitted by regressing log^- transformed dry weights on age: where W^ = dry weight (g) at age Md); Wq = dry weight (g) at hatch (the y-intercept of the log-linear regression); and G = weight-specific growth coefficient (/d). Coefficients in growth-model regressions were com- pared among regions and between cruises (months) m analysis of covariance ( ANCOVA). The ANCOVAs tested for differences in slopes (growth rates) andy- intercepts in the growth equations. When significant differences were found, a multiple range test (Stu- dent-Newman-Keuls) was applied to determine which of the growth rates differed significantly. The 558 Fishery Bulletin 97(3), 1999 mean baywide growth-in-length and growth-in- weight rates for each cruise were estimated from regressions fitted to pooled length and age data from all stations. Because larvae collected in June were <13 mm SL, growth rates of larvae collected in July were estimated from two separate regressions — one for larvae <13 mm SL that was directly comparable to the June data and one for larvae >13 mm SL. Mortality rates Age-length keys were developed to convert larval length distributions to age distributions from which mortality rates then were estimated. To derive the keys, linear regressions, based upon subsamples of otolith-aged larvae from each cruise and region, were fitted to larval age-on-length relationships. For each regression, the standard error of the estimated re- gression coefficient was used to calculate a standard normal deviate (z-statistic) from which probabilities of ages of larvae within LO-mm length classes were obtained. Six age-length keys were constructed, one for each region and cruise. This maintained the re- gion-specific integrity of size-at-age data, allowing estimation of region-specific mortality rates. Instantaneous daily mortality rates of larvae were estimated from an exponential model of decline in abundance with respect to age: N, where A'^, N. abundance ( number/m''^ 1 at age Md); estimated initial abundance (y-inter- cept of regression; number/m-); M = instantaneous mortality coefficient (/d); and t = age (d). The data were fitted to the log-linear form of the model after log_,-transformation of the abundance data. Region-specific mortality coefficients were esti- mated and compared within each cruise by analysis of covariance (ANCOVA). Mean baywide mortality rates for each cruise (month) were estimated by pool- ing abundance-at-age data from all stations and then compared in ANCOVA. When significant, ANCOVAs were followed by a multiple comparison test (Stu- dent-Newman-Keuls) to determine which mortality estimates differed significantly Two separate mor- tality rates were estimated for larvae collected in July — one for larvae <13 mm SL (<18-day-old lar- vae) that was directly comparable to the June data that included only larvae of those ages, and one for larvae >13 mm SL (>18-day-old larvae). Length-spe- cific mortality rates also were estimated, by regress- ing logp-transformed abundances of larvae on 1-mm length classes. M/G ratio and stage-specific survivorship Stage-specific survival can be estimated from the M / G ratio, where M is the instantaneous mortality rate, and G is the weight-specific growth coefficient. The M/G ratios were compared between cruises and among the three designated regions of Chesapeake Bay. The ratio M/G is an indicator of stage-specific survivorship and production potential of larval cohorts ( Houde, 1996, 1997a, 1997b). The ratio, sometimes termed the "physi- ological mortality rate," expresses a population's mor- tality per unit of individual growth (Beyer, 1989). Stage-specific survival of bay anchovy larval co- horts was estimated as [W /wj-'-w/G.^ where S = stage-specific survival = ^J^o', N^ = number of survivors at the end of a stage; Nn number alive at the beginning of a stage; W^ = dry weight of a 12-mm-SL bay anchovy larva (1000 mg); Wg = dry weight of a 3-mm-SL bay anchovy larva ( 10 g). [Note: This weight is more accurate than weight estimated for a 3-mm larva from the weight-length re- lationship, which overestimated weights of the smallest larvae]; and M/G = ratio of instantaneous mortality coef- ficient (M) and weight-specific growth coefficient (G). Survival to 12 mm SL, the largest length class fully represented in collections in each of the months, was calculated for each region by multiplying stage-spe- cific survival rate (S) by estimated abundance of the smallest fully represented length class (i.e. A^,, at 3 mm SL). Stage-specific survival rates also were esti- mated for egg to 3-day larva, 3-day to 10-day larva, and 10-day to 18-day larva. Age-specific production at a station was obtained by multiplying estimated larval density (number/m*) by the volume repre- sented by the station. Regional productions were obtained by summation of larval abundances for all stations in each region. Correlations and predictions Multiple regression analyses were applied to deter- mine if bay anchovy larval growth and mortality Rilling and Houde: Variability in growth and mortality of Anchoa mitchilll 559 rates could be related to biological or environmental factors. The four independent variables considered for inclusion were gelatinous predator biovolume (mL/m-^), zooplankton density (number/L), tempera- ture, and salinity. Mean regional estimates of inde- pendent variables were entered into the regression model. Prior to multiple regression analysis, simple correlation analyses were run to determine which independent variables might be colinear and unsuit- able for inclusion in the multiple regression model (SAS Institute, 1990). Pairs of independent variables with a correlation coefficient >0.70 were considered highly correlated and thus excluded from the mul- tiple regression analysis. significantly between June and July 1993 (^-test, P<0.0001). In each month, there was a well estab- lished pycnocline. Baywide mean temperature in- creased from 25.3°C in June to 26.6°C in July Mean salinity baywide increased from 10.1 psu in June to 15.9 psu in July (Table 1) and increased in both months between the upper and lower regions of the Bay (ANOVA, P<0.001). Mean dissolved oxygen lev- els decreased from 7.8 mg/L in June to 6.3 mg/L in July. In June, 3-m DO levels ranged from 4.1 to 12.6 mg/L ; in July they were lower, 4.6 to 8.4 mg/L. Hy- poxic and near-anoxic conditions (<2.0 mg O2/L) were most prevalent in the deep channel of the mid bay region, especially in July. Zooplankton analyses In the laboratory, zooplankton organisms were iden- tified by using a dissecting microscope. Zooplankton that were potential prey of larval anchovy were enu- merated, i.e. copepods, barnacle nauplii, gastropod veligers, bivalve veligers, cladocerans, rotifers, tintinnids, polychaete larvae, and chaetognaths. Cope- pods were categorized as adults, copepodites, and nau- plii. Densities of zooplankters were calculated as D=NIV. where D = density of organisms (number per liter); N = number of organisms in a 10-L Niskin- bottle sample: and V - sample volume ( 10 liters). Densities were weighted according to the depth range represented by each sample to ob- tain a weighted mean density in the entire wa- ter column: * 1=1 D m ; = 1 where m = depth ranges represented by each Niskin-bottle sample: and k = number of Niskin-bottle samples taken on a CTD cast (3 or 4). Results Hydrography Mean temperatures, salinities, and oxygen lev- els at 3-m depth in Chesapeake Bay differed Growth Baywide, mean growth rate of anchovy larvae in- creased from 0.59 mm/d in June to 0.72 mm/d in July (ANCOVA, P<0.001) (Fig. 2). When the analysis in- cluded only larvae <13 mm SL, an even higher bay- wide growth rate was estimated in July (0.78 mm/d) than in June (0.59 mm/d) (ANCOVA, P<0. 001). Esti- mated lengths at 15 days after hatching were 11.65 mm SL in June and 13.72 mm SL in July. Growth rates did not differ significantly among regions in June or July (ANCOVA, P>0.05) (Fig. 3). The regional estimates of growth-in-length rates ranged from 0.53 mm/d in the lower bay during June to 0.78 mm/d in the upper bay in July. Although not significantly different, the highest regional growth rates for each month were in the upper bay. The 30 25 20 J, 15 I .0 JuneL = 2 80+0 59d D July r: = 0 84 n= 128 S^ = 0 01 n ■ Jul>' L = 2 92+0,72d n n^/ r= = 0 89 n = 296 S, = 0 02 jr^am a 1° m^' n iSiiil^ vr D^jllR; jP^ 1 1 10 15 20 Age (days) 25 30 35 Figure 2 Baywide linear growth models for bay anchovy larvae in Chesa- peake Bay, June and July 1993. L = standard length (mm), d = age in days, estimated from otolith-increment analysis. ♦ = June, = July 560 Fishery Bulletin 97(3), 1999 to June L = 280 + 061d n=55 S|, = 0 04 r^ = 0 83 Upper Bay L=3.22+0.57d n = 42 S^ = 0 04 Mid Bay July L = 2 98 + 0 72d Lower Bay n=117 S^ = 0.02 ^^^ r2 = 0 87 A^^^^ -J -irR 10 15 20 25 30 Age (days) Figure 3 Linear growth models for bay anchovy larvae in upper, mid, and lower Chesapeake Bay regions, June and July 1993. L = standard length (mmi, d = age in days, estimated from otolith-increment analysis. v-intercepts of the linear regressions ranged from 2.33 mm to 3.22 mm and did not differ significantly among regions or between June and July (ANCOVA, P>0.05). Growth in weight was rapid for surviving larvae. Baywide, estimated weight-specific gi-owth rates G {Id) for larvae <13 mm SL (Table 1) increased from 0.26 in June (29.7'7r/d) to 0.35 in July (41.9%^) (ANCOVA, P<0.05 ). For the <13 mm larvae, the high- est regional G was 0.40 in the upper bay in July (49.2%/d), and the lowest G was 0.25 in the mid bay in June (28.4%/d). In July, an estimated baywide G for larvae >13 mm SL was only 0.11 (11.6%/d), a rate much lower than that of smaller larvae. Mortality Larvae experienced high mortality rates. Baywide instantaneous mortality coefficients M (/dl for lar- vae in all age classes declined significantly from 0.41 (33.6%/d) in June to 0.23 (20.5%/d) in July (ANCOVA, P<0.0001) (Fig. 4). When the data analysis included only larvae <1 8-day s old (the oldest age represented in June) estimated July M = 0.22 (19.7%/d), a rate still significantly lower than the June rate (ANCOVA, P<0.001). Larval mortality rates differed significantly among regions in June and July ( ANCOVA, P<0.0001 ). High- est mortality rates in each month were in the lower bay. Regional daily rates ranged from 0.14 (13.1'^/d) to 0.54 (Al.l'AIA) (Fig. 5). In June, the mid bay mor- tality rate was significantly lower than upper and lower bay rates. In July, when all regional mortality rates differed significantly, the lower bay had the highest mortality rate and the upper bay had the lowest rate (Student-Neumann-Keuls test). Length-specific mortality rates (/mm ) declined dra- matically as larval length increased (Fig. 6). In June, the length-specific mortality rates ranged from 0.45/ mm for 2-5 mm SL larvae to 0.07/mm for 10-13 mm SL larvae. In July, the rates were lower, ranging from 0.33/mm for 2-5 mm SL larvae to 0.05/mm for 10-13 mm SL larvae. The highest regional length-specific rate, 0.64/mm, occurred in the mid bay during July for the 2-5 mm SL class. M/G ratio, production, and survival The baywide MIG ratio declined from 1.59 in June to 0.67 in July (Fig. 7), indicating a nearly 70-fold higher survival and biomass production of cohorts Rilling and Houde; Variability in growth and mortality of Anchoa mitchtlli 561 Table 1 Summarized data, Chesapeake Bay cruises, 19-22 June and 23-30 July 1993. Data include anchovy weight-specific growth rates (/d), mean biovolumes (mL/m^') of gelatinous predators Mnemiopsis leidyi and Chrysaora quinquecirrha. mean densities (organ- isms/L) of zooplankton (ZOOPi, mean regional temperatures (T) and salinities (S) at 3-m depth. Identical superscripts indicate no significant difference (P>0.05, Tukey's multiple range test) among regions or between months. SE = standard error. Region T(°C) S (psu) ZOOP (/L) Vol. of M. leidyi (mL/m^) Vol. of C. quinquecirrha (mL/m2) Weight-specific growth rate G (/d) June Upper bay 25.4'" 5.8° 326.8° 41.4° 0.0° 0.28° SE 116.7 22.1 — 0.01 Mid bay 25.3° 9.7' 23.1'' 999.5'' 0.0° 0.25° SE 7.2 211.3 — 0.01 Lower bay 25.3° 14.7c 68.6* 516.8° 0.0° 0.25" SE 21.7 134.4 — 0.02 July Upper bay 26.8° 8.9° 442.5° 229.6" 30.7" 0.40° SE 91.3 65.2 12.1 0.02 Mid bay 26.8" 12.6'' 162.5'' 204.5° 36.4" 0.37° SE 35.5 44.7 7.9 0.03 Lower bay 26.2" 26.2c 277.2" 12.1" 22.0" 0.28" SE 40.0 7.2 11.8 0.02 Baywide June 25.3° 10.1" 138.5° .560.7° 0.0° 0.26" SE 43.9 104.2 — 0.01 July 26.6* is.g* 289.0'' 139.1" 29.5'' 0.35'' SE 27.4 6.1 in the 3-12 mm SL size range during July. The in- crease in gi-owth and decHne in mortahty rates be- tween June and July accounted for the drops in Ml G ratio. On average, larval cohorts lost biomass in June (M/G>1.0), but gained biomass in July [M / G <1.0). In June, cohorts at 12 mm SL supported only 6.6% of the biomass present at 3 mm SL, whereas, in July, cohorts at 12 mm SL supported 457.19^ of their 3-mm-SL biomass. Each regional MIG ratio also declined between June and July. Predicted abundance of larval daily cohorts at 12 mm SL, based upon the MIG ratios and estimated regional abundance-at-age data, was highest in the upper bay in June but shifted to the mid and lower bay in July (Table 2). Stage-specific mortality rates, estimated from declines in abundances, were high- est for the youngest stages, and declined with increas- ing age (Table 3). In this analysis, abundances were estimated for cohorts at 18 days after hatching, when regional mean lengths ranged from 12.4 to 16.4 mm SL. Despite highest regional mortality, daily cohorts from the lower bay in July produced the most 18- day-old larvae (1.5 x 10*). The daily production of 18-day-old larvae in the lower bay was 3.4-4.6 times higher than in other regions in July, and from 7 to 50 times higher than in other regions in June. Cumulative mortalities from egg to 18-day old lar- val stage were lowest in the mid bay in June but lowest in the upper bay in July. Interestingly, those regions had experienced the greatest egg to 3-day- old larval mortalities, which then were followed by low mortality in older larvae (Table 3). Predators and prey There was a significant between-cruises difference in mean combined biovolumes of two common gelati- nous predators of bay anchovy eggs and larvae, the ctenophore Mnemiopsis leidyi, and the scyphomedusa Chrysaora quinquecirrha, (t-test,P<0.00l) (Table 1). Mean biovolumes of the ctenophore shifted region- ally and declined by a factor of four in July. The scyphomedusan did not occur in June and had a mean biovolume of 29.5 niL/ni" in July, and there was no indication of regional differences (AN OVA P>0.05) (Table 1). 562 Fishery Bulletin 97(3), 1999 June Log^ Abund = 2.69 - 0 41'Age r2 = b.96 S =0.02 o -J July Log^ Abund =0 46 -0 23'Age r2 = 0 91 S^ = 0,01 ^^.^» s. ♦♦> ^^ ♦ 2 0 ■ -2 -4 -6 0 5 10 15 20 25 30 Age (days) Figure 4 Age-specific survival curves for bay anchovy larvae in Chesapeake Bay. June and July 1993. The age-specific abundance estimates for each day class were derived from age-length keys. Table 2 Predicted cohort survivorships of bay anchovy larvae in the upper, mid, and lower Chesapeake Bay in June and July 1993. N^, = estimated number of smallest fully repre- sented length class of larvae (3 mm SLl. M/G = ratio of instantaneous mortality rate (/d ) to weight- specific growth rate (/d), survival rate (S) = IW/W,,]-'""''' where W = dry weight of a 12-mm-SL larva (1000 //g), W„ = dry weight of a 3-mm-SL larva (10 mgl, and apparent survivorship at 12 mm = SxNg. Apparent Ng survivorship (3-mm-SL Survival at Region larvae) M/G rate 12 mm SL June Upper bay Mid bay Lower bay July Upper bay Mid bay 2.59 X IQi" 2.49 X lO* 7.28 X 10' .5.95 X 10» 2.08 X 10'" 1.72 0.87 2.17 0.35 0.62 0.0004 0.0180 0.00005 0.200 0.058 Lower bay 11.72x10'" 1.19 0.004 1,04 X 10' 4.48 X 10« 3.64 X 10'' 1.19 X lO" 1.21 X 10» 4.69 X 10** Baywide, the mean density of zooplankton that are potential prey of anchovy larvae doubled between June and July (f-test, P<0.05) (Table 1). Mean density of zooplankton was highest in the upper bay in June (ANOVA, P<0.05) and was higher in the upper and lower bay than in the mid bay in July (ANOVA, P<0.05). Nauplii of the copepod Acartia tonsa were the single most abundant zooplankter col- lected. Tintinnids, rotifers, and the cyclopoid copepod Oithona sp. also were common. Mean density of copepod nauplii, a major prey of bay anchovy larvae, increased from 36.9/L in June to 110.5/L in July (i-test, P<0.05). Correlations At the regional level, few correlations were judged to be significant at the a = 0.05 level between biological and environmental variables (Table 4). The low degrees of freedom (/!=6) and corresponding low power made it difficult to reject null hypotheses. Several coefficients were high enough to suggest possible correlations. Anchovy larval abundances were positively cor- related with egg abundances {r-+0.96, P<0.01) and negatively correlated with gelatinous predator biovolumes(r=-0.87,P<0.05) (Fig. 8A). Larval growth rate was positively correlated with temperature (r=-i-0.94, P<0.01) (Fig. 8B) and possibly related to zooplankton density (r=-i-0.72, P<0.11). Although not significant at a = 0.05, anchovy egg abundances and zooplankton den- sities both may have been negatively correlated with gelatinous predator biovolumes (r =-0.76, P=0.08). Predicting larval growth and mortality Regional variability in anchovy larval growth-in- length rate for the combined June and July cruises was explained reasonably well by a two-variable re- gression model that included temperature and zoop- lankton density (r-=0.93): g = 0.32 + 0.05X, + 0.004X^, where g = larval growth rate (mm/d); X, = log^, zooplankton density (organisms/L); and X,, = temperature (°C). Larval growth rates increased with increasing tem- peratures and zooplankton densities. Temperature accounted for more of the variability in growth rate than did zooplankton density. Observed and model- Rilling and Houde: Variability in growth and mortality of Anchoa mitchi/li 563 June July 4 2 Upper Bay \^ Log, Abund = 4.72 - 0 48*Age ^^> r= = 0 93 Sj = 0 04 n = 15 Upper Bay Log, Abund = -0 85 - 0 14*Age r! = 0.70 Si, = 0 02 n = 22 0 -2 ^ -4 rM e -6 i 2j 77^**^***.^**u^^ Mid Bay Log, Abund = -0 14 -0 21*Age r= = 0 83 Sj = 0,03 n = 16 • Mid Bay Log, Abund = -0 98 - 0.23'Age r' = 0.84 Sfc = 0.02 n = 26 g 0 1 -2 § -4- ^^*^~tr»***^^^ *'^^^^>**^ii^. 2 4 2 0 -2 -4 Lower Bay I. Log, Abund = 323-0 54*Age \,^ rJ = 0 93 Sb = 0 04 /! = /« Lower Bay Log, Abund = 3 09 - 0 34'Age r! = 0 91 Sfc = 0 02 n = /P ^*^^ * ^***D(M/d) Cumulative mortality B^D Apparent survival June Upper 8.87 X lO'i 2.59 X 10"' 1.18 4.42 X 10" 0.58 2.16 X 10' 0.38 7.09 0.0008 Mid 10.44 X 10" 2.49 X 10» 2.78 1.06 • 10" 0.12 6.52 X 106 0.35 3,64 0.0263 Lower 23.28 X 10" 7.28 X 10^ 1.92 1.42 X 10" 0.56 2.45 X 1015 0.51 8.00 0,0003 July Upper 3.21 X lO" 5.95 X 10'- 2.09 4.14 < 10' 0,38 3.26 X 10- 0.03 2.90 0,055 Mid 1.53xl0'2 2.08 X IQi" 1.43 1.33x108 0.72 4.39 X 10^ 0.14 6.16 0.002 Lower 6.58 X 1012 11.72 V 101" 1.34 9.29 X 108 0.69 1..50 X 10" 0.22 6.66 0.001 predicted values of mean regional larval growth rates differed by no more than 6'7f. No satisfactory multiple regression model could be fit for larval mortality. Discussion Cohorts of bay anchovy larvae in Chesapeake Bay do not experience uniform growth and mortality. 564 Fishery Bulletin 97(3), 1999 Growth rates were temporally variable, and mortal- ity rates were both spatially and temporally variable. It was clear that variability in these rates could sig- nificantly alter production and recruitment poten- tials. Baywide, larval growth rates were higher in July than in June. The larval growth rates that we report are higher and more variable than rates re- ported previously for bay anchovy in the laboratory o E s. 0.75 05 0.25 075 05 025 June 2-5 6-9 10-13 >14 July 2-5 6-9 10-13 Length (mm) >14 Figure 6 Baywide mean length-specific mortality rates (/mm) of bay anchovy larvae in four length classes during June and Julv 1993. Error bars are =1 SE. * = no data. and in most field studies (Table 5). Gallagher et al. ( 1983) did report equivalent and higher growth rates (0.59-0.93 mm/d) in the Patuxent River tributary of Chesapeake Bay. Temporal and spatial variability in anchovy lar- val growth rates was related to both zooplankton density and temperature. The mean baywide growth rate of larvae increased from 0.59 mm/d in June to 0.72 mm/d in July, corresponding to increases in mean water temperatures and mean copepod nau- plii densities. The increase in larval growth rate be- tween months corresponded to a coincident 1.3°C increase in mean temperature at 3-m depth and to a major increase in zooplankton abundance. On the basis of laboratory experiments, Houde (1978) pre- dicted minimal prey concentration for lO^c sui-vival of bay anchovy larvae at 26" C to be 107 copepod nau- a s 1 o M G M/G M G M/G June July Figure 7 Baywide daily instantaneous mortality rate (M), weight- specific growth rate (G). and M/G ratios for bay anchovy larvae in Chesapeake Bay, June and July 1993. Table 4 Correlation coefficients at the regional level (n=6) for data collected in Chesapeake Bay, June and July 1993. The seven variables include mean regional instantaneous larval mortality rate (M. /d), larval growth rate (g, mm/d), egg and larval abundances (number/ m^), log^.-combined gelatinous predator biovolumes (mL/m^) of Mnemiopsis leidyi and Chrysaora quinquecirrha, zooplankton density lorganisms/Ll, and temperature and salinity at 3-m depth. * = P<0.05. ** = P<0.01. In each case, n = 6, r^^^ = 0.81. M g Egg abundance Larval abundance Jelly biovolume Zooplankton density Temperature (°C) g -0.68,5 Egg abundance 0.364 0.0,56 Larval abundance 0.434 0.165 0.955** Jelly biovolume -0.287 -0.380 -0.765 -0.868* Zooplankton density -0.039 0.721 0.251 0.477 -0.761 Temperature -0.771 0.938** -0.027 0.037 -0.170 0.555 Salinity 0.111 0.091 0.524 0.521 -0.132 -0.072 0.067 Rilling and Houde: Variability in growth and mortality of Anchoa mitchil/i 565 plii/L. In the present study, nauplii densi- ties often exceeded that level in July, but levels were lower and potentially limiting during June, especially in the mid bay (mean = 9.7 nauplii/L) and lower bay (mean=17.7 nauplii/L). During each month, fastest gi'owth rates were estimated for larvae from the upper bay where copepod nauplii were most abundant (means=84.8 nauplii/L in June and 157.5 nauplii/L in July). Lowest growth was estimated during June in the mid and lower bay, where copepod nauplii densities and mean temperatures were lowest. Although temperature alone may exercise an important control over lar\'al growth, (Houde, 1989a: Pepin, 1991), a combination of factors, including prey availability, effects of body size, or growth- rate dependent mortality, may operate to control growth rates and survival poten- tial (Bailey and Houde, 1989; Heath, 1992; Leggett and DeBlois, 1994). In the case of bay anchovy, all of these factors may oper- ate, but temperature and prey level appar- ently predominate. There also was an effect of body size; growth rate of larvae >13 mm declined. In a synthesis analysis, Houde (1997b) reported that average weight-spe- cific growth coefficients of bay anchovy de- clined progressively from 0.573 ( 77.3T/d ) in newly hatched lawae to 0.065 l6.7'J/d) in near-metamorphosis individuals. Our baywide instantaneous daily mor- tality rates decreased from 0.41 in June to 0.23 in July 1993. This decline was co- incident with increasing growth rate, sug- gesting that cohorts of rapidly growing larvae in July might have been less vul- nerable to size-selective or growth-rate dependent predation. From length-specific and age-specific analyses of mortality, it was clear that mortality rates were great- est for the smallest and youngest lai"vae (Table 3). Houde (1997b), analyzed the accumulated data on bay anchovy larvae from Chesapeake Bay and demonstrated that mortality rate (A/) declined predictably with respect to body weight raised to the -0.318 power In the present study, mortality from the egg to 3-day-old larval stage was 2 to 7 times higher than mortality from the 3- to 10-day-old lar- val stage. Interestingly, mean baywide mortality rates for the egg to 3-day-old stage (yolksac and first- feeding larvae) were similar in June and July (82'7f/d in June, 797c/d in July), but mortality rates for the o O Log Jellyfish biovolume (mL/m') 08- B 075- Growth = -2 59 + 0 12»Temperature r = = 0 88 '^ 07h ■ -^ 0.65- ^/^^"^ 06. -^^^ ^^^^B 0 55- ^^^^ OV < 1 1 1 1 1 1 1 1 25 25,5 26 26.5 27 Temperature (°C) Figure 8 Relationship between I A) mean regional bay anchovy larval abundance and gelatinous predator biovolumes and ( B i mean regional bay anchovy larval growth rates and temperature at 3-m depth in Chesapeake Bay. June and Julv 1993. 10 to 18-day-old larval stage were considerably lower in July (34'/f/d in June, I27c/d in July), implying that conditions had become more favorable for feeding- stage larvae in July. Baywide cumulative mortality rates for egg to 18-day-old larvae indicated that <0.1'7f of a daily cohort survived to 18 days after hatching in June and that -1.6% survived to 18 days in July Predation is a major cause of mortality in the early life of marine fishes (Leggett, 1986; Bailey and Houde, 1989; Leggett and DeBlois, 1994) and may 566 Fishery Bulletin 97(3), 1999 Table 5 Reported growth and mortality rates of bay anchovy larvae in field studies. Location Growth 1 mm/d ) Mortality (M/d) Source Patuxent River, MD 0.59-0.93 Gallagher et al. (1983) Newport River Estuary, NC 0.25-0.51 Fives etal. (1986) Biscayne Bay, FL 0.43-0.56 0.30-0.45 Leak and Houde ( 1987 ) Mesocosnis in Chesapeake Bay, MD 0.39-0.61 0.08-0.23 Cowan and Houde (1990) Great South Bay N\' 0.52-0.59 0.32-0.89 Castro and Cowen(1991) Chesapeake Bay, MD (yolksac larvae) 0.41-4.24 Dorsey et al. 11996) Chesapeake Bay, MD 0.23-1.20 MacGregor and Houde (19961 Chesapeake Bay, MD 0.5.3-0.78 0.14-0.54 This study, range of regional rates be the major agent of mortality operating on bay anchovy in Chesapeake Bay. The gelatinous zoo- plankters M. leidyi anci C. quinquecirrha, are known to be important predators on eggs and larvae of bay anchovy (Feigenbaum and Kelly, 1984; Monteleone and Duguay, 1988; Cowan and Houde, 1993; Purcell et al., 1994). In Chesapeake Bay, the peak periods of bay anchovy spawning and gelatinous zooplankton abundance overlap during June and July ( Cowan and Houde, 1993; Purcell et al., 1994), facilitating the predator-prey interaction. We found gelatinous predator biovolumes to be significantly higher in June, when only M. leidyi was present, than in July when both species occurred. Chrysao/'a quinque- cirrha, a potentially more powerful predator on an- chovy eggs and larvae than M. leidyi (Purcell et al., 1994), occurred only in July, but at low mean biovolumes that were uniform in the three Bay re- gions. It is worth noting that C. quinquecirrha is also a predator on M. leidyi (Purcell and Cowan, 1995), resulting in a predator-prey interaction that poten- tially has a sparing effect on anchovy eggs and lar- vae (Cowan and Houde, 1992). The large biovolumes of gelatinous predators prob- ably contributed to the greater mortality rates in June compared with July. However, the six mortal- ity coefficients from the regional estimates were not significantly correlated with gelatinous predator biovolumes (Table 4), despite the strong, negative, linear relationship between anchovy larval abun- dance and gelatinous predator biovolume (r^=0.75; Fig. 8A) for the combined June and July regional data. This negative correlation may have been a con- sequence of predation, but it also could have been generated by lower egg production of anchovy in ar- eas where jellyfish were abundant, as Dorsey et al. (1996) hypothesized in site-specific studies of bay anchovy egg and yolksac larval mortality. Although we cannot conclude unequivocally that gelatinous predators accounted for high mortality rates, it is likely that they were significant consumers of an- chovy larvae. Abundance data illustrated in survival curves (Figs. 4 and 5) showed several modes, which sug- gested that cohort-specific mortality might be vari- able on shorter time scales than we studied and might be changing as a function of ontogeny, age, or size. Other factors also could have biased our mortality estimates or contributed to regional variability in rates, for example, pulses of spawning that produce variable initial abundances of daily cohorts or immi- gration and emigration of larvae into and out of a region. Because mortality is size-specific, the pres- ence of larger larvae in July could have led to a lower estimate of mortality rate during that period. But, when only larvae of equivalent lengths (i.e. <13 mm SL) were analyzed, estimated mortality rates re- mained nearly twice as high in June (M=0.41) as in July (M=0.22). If some larvae were being advected up the bay, as Dovel (1971) had hypothesized, this process could have contributed to biased estimates of higher mortality rates in the lower bay. Despite a high cumulative mortality rate, the lower bay in July had the highest production of larvae sur- viving to 18 days after hatching for the June-July 1993 period (Table 3). This result is due to high spawning activity and initial concentrations of lar- vae in the lower bay ( Rilling and Houde, manuscript in review), a high growth rate of larvae, and, impor- tantly, the relatively large volume of water in the lower bay, which supported a large contingent of anchovy larvae. Survival and recruitment potential of anchovy co- horts were responsive to variability in both mortal- ity rates and growth rates that they experienced. The ratio M/G, an index of stage-specific mortality, is an important indicator of comparative production and survival potential during early life (Houde, 1996, Rilling and Houde: Variability in growth and mortality of Anchoa mitchilli 567 1997a). The baywide M IG ratio for bay anchovy lar- vae decUned from 1.59 in June to 0.67 in July. The low MIG ratio in July is a reflection of the coinci- dent decline in larval mortality rate and increase in growth rate that occurred between June and July. The difference in MIG ratios between months im- plies a 70-fold higher survival potential through the larval stage for July-hatched cohorts. MIG ratios <1.0, signifying high regional produc- tion potential, were observed in the mid bay in June and in the mid and upper bay in July. Larvae in those regions tended to have lowest mortality rates. An abundance of large larvae generally indicates higher survival rates of cohorts, but size-selective mortal- ity or transport (or both) of larvae into a region (Fortier and Leggett, 1982, 1985; Norcross and Shaw, 1984; Boehlert and Mundy, 1988) could have contrib- uted to the relative abundances of large larvae and low MIG ratios. In the Patuxent River subestuary of Chesapeake Bay, progressive increases in larval length upriver were reported by Loos and Perry (1991), who hypothesized (with supporting evidence) that transport of larvae was primarily responsible. Similarly, MacGregor and Houde (1996) reported a gradient in bay anchovy larval size on a cross-Bay transect that was repetitively sampled; smallest lar- vae were found offshore and largest larvae, inshore. In the present study, selective up-bay transport of larx'ae could have acted to reduce the MIG ratio in the upper bay between June and July, but we cannot confirm it. In summary, mortality rates of bay anchovy early- life stages were both temporally and regionally vari- able at one-month temporal and at 60-km spatial scales in Chesapeake Bay. Growth rates showed strong temporal variation but no significant regional differences. Stage-specific survival, which depended upon both mortality and growth rates, was both size- specific and growth-rate dependent. We found no obvious indication of density-dependent mortality (i.e. no correlations between egg or larval abundances and mortality or growth rates), although recent in- dividual-based modeling suggests that density-de- pendence in early-life could be an important regula- tor of bay anchovy recruitments in Chesapeake Bay (Wang et al., 1997). The lower Chesapeake Bay in July was the major source of potential recruits in 1993. Temperature, zooplankton prey, and gelatinous predators all are believed to have contributed to tem- poral and regional differences in growth and mortal- ity of larvae. Further research is needed to define scales and patterns of processes that control vari- ability in production and recruitment of bay anchovy. This will require coupled biophysical studies and development of models that, up to now, have essen- tially emphasized only biology. Acknowledgments Research was supported by National Science Foun- dation grants OCE-92-03307 and OCE95-21512 to E. D. Houde and by NSF Land Margin Ecosystem grant DEB94-12113 to W. R. Boynton et al. We thank the officers and crew ofRVHenlopen for capable re- search-vessel support and numerous colleagues and students who assisted in the research. We especially thank S. Leach and L. Fernandez for assistance with illustrations and preparation of the manuscript. Literature cited Bailey, K. M.. and E. D. Houde. 1989. Predation on early developmental stages of marine fishes and the recruitment problem. Adv. Mar. Biol. 25:1-83. Baird, D., and R. E. Ulanowicz. 1989. The seasonal dynamics of the Chesapeake Bay ecosystem. Ecol. Monogr. .59(4):329-.364. Beyer. J. E. 1989. Recruitment stability and survival-simple size-spe- cific theory with examples from the early life dynamics of marine fish. Dana 7;4.'5-147. Boehlert, G. W., and B. C. Mundy. 1988. Roles of behavior and physical factors in larval and juvenile fish recruitment to estuarine nursery areas. Am. Fish. Soc. -Symp. 3:,51-67. Bolz, G. R., and B. R. Burns. 1996. Age and growth of larval Atlantic herring, Clupea harengus: a comparative study. Fish. Bull. 94:387-397. Castro, L. R., and R. K. Cowen. 1991. Environmental factors affecting the early life history of bay anchovy Anc/ioa mitchilli in Great South Bay, New York. Mar. Ecol. Prog. Ser. 76:235-247. Cowan, J. H., Jr., and E. D. Houde. 1990. Growth and sur\'ival of bay anchovy .A^c/ioa mitchilli larvae in mesocosm enclosures. Mar. Ecol. Prog. Ser. 68:47-57. 1992. Size-dependent predation on marine fish larvae by ctenophores, scyphomedusae and planktivorous fish. Fish. Oceanogr. 1:113-126. 1993. Relative predation potentials of scyphomedusae, ctenophores and planktivorous fish on ichthyoplankton in Chesapeake Bay. Mar. Ecol. Prog. Ser. 95:55-65. Gushing, D. H. 1975. Marine ecology and fisheries. Cambridge Univ. Press, Cambridge, U.K. 278 p. Dalton, P. D. 1987. Ecology of bay anchovy, Anchoa mitchilli, eggs and larvae in the mid-Chesapeake Bay M.S. thesis. Univ of Maryland. College Park, MD, 104 p. Dorsey, S, E., E, D. Houde, and J. C. Gamble. 1996. Cohort abundances and daily variability in mortal- ity of eggs and yolk-sac larvae of bay anchovy, Anchoa mitchilli. in Chesapeake Bay Fish. Bull. 94:257-267. Dovel, W. L. 1971. Fish eggs and larvae of the upper Chesapeake Bay. Natural Resources Inst., Univ. of Maryland Special Report 4. 71 p. Feigenbaum, D., and M. Kelly. 1984. Changes in the lower Chesapeake Bay food chain in 568 Fishery Bulletin 97(3), 1999 presence of the sea nettle Chrysaora quinquecirrha (schyphomedusa). Mar. Ecol. Prog. Sen 19:39-47. Fives, J. M., S. M. Warlen. and D. E. Hoss. 1986. Aging and growth of larval bay anchovy Anchoa mitchillt, from the Newport River Estuary, North Carolina. Estuaries 9;362-367. Fortier, L., and W. C. Leggett. 1982. Fickian transport and the dispersal offish larvae in estuaries. Can. J. Fish. Aquat. Sci. 39:1150-1163. 1985. Adrift study of larval fish survival. Mar. Ecol. Frog. Ser. 25:245-257. ' Gallagher, R. P., M. F. Hirshfield, and E. S. Perry. 1983. Age, growth, and mortality estimates for larval bay anchovy An c/ioa mitchiUi in the Patuxent River. Benedict Estuarine Research Laboratory publication, Division of Environmental Research; Academy of Natural Sciences of Philadelphia, Philadelphia. PA, 34 p. Hartman, K. J., and S. B. Brandt. 1995. Comparative energetics and the development of bioenergetics models for sympatric estuarine piscivores. Can. J. Fish. Aquat. So. 52:1647-1666. Heath, M. R. 1992. Field investigations of the early life stages of marine fish. Adv. Mar Biol. 28:3-174. Hildebrand, S. F., and W. C. Schroeder. 1928. Fishes of Chesapeake Bay. Bull. U.S. Bur Fish. 43:108-110. Houde, E. D. 1 978. Critical food concentrations for larvae of three species of subtropical marine fishes. Bull. Mar Sci. 28:39.5^11. 1987. Fish early life dynamics and recruitment variability. Am. Fish. Soc. Symp. 2:17-29. 1989a. Comparative growth, mortality, and energetics of marine fish larvae: temperature and implied latitudinal effects. Fish. Bull. 87:471-495. 1989b. Subtleties and episodes in the early life of fishes. .J. Fish Biol. 35 (suppl. A):29-38. 1996. Evaluating stage-specific survival during the early life of fish. In Y. Watanabe, Y. Yamashita and Y Oozeki (eds.). Survival strategies in early life stages of marine resources, p. 51-66. Balkema, Rotterdam. 1997a. Patterns and consequences of selective processes in teleost early life histories. In R. C. Chambers and E. A. Trippel (eds.). Early life history and recruitment of fish populations, p. 173-196. Chapman and Hall, London. 1997b. Patterns and trends in larval-stage growth and mortality of teleost fish. J, Fish Biol. 51(suppl. A):52-83. Houde, E. D., and C. E. Zastrow. 1991. Bay anchovy. In S. L. Funderburk, J. A. Mihursky, S. J. Jordan, and D. Riley (eds.). Habitat requirements for Chesapeake Bay living resources, 2nd ed., p. 8-1 to 8- 14. Living Resources Subcommittee. Chesapeake Bay Program, Annapolis, MD. Leak, J. C. 1986. The relationship of standard length and otolith di- ameter in larval bay anchovy, Anc/ioa mitchilli, a shrink- age estimator J. Exp. Mar Biol Ecol. 95:167-172. Leak, J. C, and E. D. Houde. 1987. Cohort growth and survival of bay anchovy Anc/ioa mitchiUi larvae in Biscayne Bay, Florida. Mar Ecol. Prog. Ser 37:109-122. Leggett, W. C. 1986. The dependence of fish larval survival on food and predator densities. In The role of freshwater outflow in coastal marine ecosystems, p. 118-137. NATOASI series, vol. G7. Leggett, W. C, and E. DeBlois. 1994. Recruitment in marine fishes: is it regulated by star- vation and predation in the egg and larval stages? Nether- lands J. Sea Res. 32:119-134. Loos, J. J., and E. S. Perrj'. 1991. Larval migration and mortality rates of bay anchovy in the Patuxent River. U.S. Dep. Commer, NOAA Tech- nical Report NMFS 95:65-76. Luo, J., and J. A. Musick. 1991. Reproductive biology of the bay anchovy in Chesa- peake Bay Trans. Am. Fish. Soc. 120:701-710. MacGregor, J. M. 1994. Temporal and spatial variability in cross-bay distri- bution and abundance of bay anchovy ^Anchoa mitchilli) eggs and larvae. M.S. thesis, Univ. Maryland, College Park, MD, 138 p. MacGregor, J. M., and E. D. Houde. 1996. Onshore-offshore pattern and variability in distribu- tion and abundance of bay anchovy Anc/!oa mitchilli eggs and larvae in Chesapeake Bay. Mar Ecol. Prog. Ser. 138:5-25. Monteleone, D. M., and L. E. Duguay. 1988. Laboratory studies of predation by the ctenophore Mnemiopsis leidyi on the early stages in the life history of the bay anchovy, Anchoa mitchilli. J. Plankton Res. 10:359-372. Newberger, T. A., and E. D. Houde. 1995. Population biology of bay anchovy, Anc/ioa mitchilli, in mid-Chesapeake Bay. Mar Ecol. Prog. Ser. 116:25-37. Norcross, B. L., and R. F. Shaw. 1984. Oceanic and estuarine transport offish eggs and lar- vae: a review. Trans. Am. Fish. Soc. 113:153-165. Olney, J. E. 1983. Eggs and early larvae of the bay anchovy, Anchoa mitchilli. and the weakfish, Cynoscion regalis. in lower Chesapeake Bay with notes on associated ichthyo- plankton. Estuaries 6(1 ):20-35. Pepin, P. 1991. Effect of temperature and size on development, mor- tality, and survival rates of the pelagic early life history stages of marine fish. Can. .J. Fish. Aquat. Sci. 48:503- 518. PurcelL J. E. and J. H. Cowan, Jr. 1995. Predation by the scyphomedusan Chrysaora quinquecirrha on Mnemiopsis leidyi ctenophores. Mar Ecol. Prog. .Ser 129:63-70, PurceU, J. E., D. A. Nemazie, S. E. Dorsey, E. D. Houde, and J. C. Gamble. 1994. Predation mortality of bay anchovy ^Anchoa mitchilli) eggs and larvae due to scyphomedusae and ctenophores in Chesapeake Bay Mar Ecol. Prog. Ser 114:47-58. SAS Institute. 1990. SAS/STAT user's guide, version 6, fourth ed, vols. I and 11. SAS Institute, Cary NC. 245 p. Secor, D. H., J. M. Dean, and E. H. Laban. 1991. Manual for otolith removal and preparation for mi- crostructure examination. LIniv. South Carolina, Colum- bia, SC, 85 p. Theilacker, G. H. 1980. Changes in body measurements of larval northern anchovy, Engraulis mordax. and other fishes due to han- dling and preservation. Fish. Bull. 78(3):68.5-692. Wang, S.-B., J. H. Cowan Jr., K. A. Rose, and E. D. Houde. 1997. Individual-based modelling of recruitment variabil- ity and biomass production of bay anchovy in mid-Chesa- peake Bay J. Fish Biol. 51(suppl. A): 101-120. Rilling and Houde: Variability in growth and mortality o\ Anchoa mitchilli 569 Wang, S,-B., and E. D. Houde. Zastrow, C. E., E. D. Houde, and L. G. Morin. 1995. Distribution, relative abundance, biomass and pro- 1991. Spawning, fecundity, hatch-date frequency and duction of bay anchovy A/ic/ioa mitchilli in the Chesapeake young-of-the-year growth of bay anchovy, A?ic/ioa mitchilli Bay. Mar. Ecol. Prog. Sen 121:27-38. in mid-Chesapeake Bay Mar. Ecol. Prog. Ser. 73:161-171. 570 Abstract. -Size at 509c maturity is commonly evaluated for wild popula- tions, but the uncertainty involved in such computation has been frequently overlooked in the application to marine fisheries. Here we evaluate three pro- cedures to obtain a confidence interval for size at 50% maturity, and in gen- eral for P% maturity: Fieller's analyti- cal method, nonparametric bootstrap, and a Monte Carlo algorithm. The three ' methods are compared in estimating size at SC* maturity t/jf,,-! by using simulated data from an age-structured population, with von Bertalanffy growth and constant natural mortality, for sample sizes of 500 to 10,000 indi- viduals. Performance was assessed by using four criteria: 1 ) the proportion of times that the confidence interval did contain the true and known size at 50'7( maturity, 2i bias in estimating /jq';- 31 length and 4i shape of the confidence interval around /-|j,.. Judging from cri- teria 2-4. the three methods performed equally well, but in criterion 1, the Monte Carlo method outperformed the bootstrap and Fieller methods with a frequency remaining very close to the nominal SS'/r at all sample sizes. The Monte Carlo method was also robust to variations in natural mortality rate (M), although with lengthier and more asymmetric confidence intervals as M increased. This method was applied to two sets of real data. First, we used data from the squat lobster Pleuron- codes monodon with several levels of proportion mature, so that a confidence interval for the whole maturity curve could be outlined. Second, we compared two samples of the anchovy Engraulis ringens from different localities in cen- tral Chile to test the hypothesis that they differed in size at 50% maturity and concluded that they were not sta- tisticallv different. Estimation of size at sexual maturity: an evaluation of analytical and resampling procedures Ruben Roa Departamento de Oceanografia Universidad de Concepcion Casilla 160-C, Concepcion, Chile E-mail address rroa a udec cl Billy Ernst School of Fisheries, WH- 10 University ol Washington, Seattle, Washington 98195 Fabian Tapia Departamento de Oceanografia Universidad de Concepcion Casilla 160-C, Concepcion, Chile Manuscript accepted 28 August 1998. Fi.sh. Bull. 97:570-580 11999). Size at 50% maturity (/jg^,) is com- monly evaluated for wild popula- tions as a point of biological refer- ence (see Table 1 ). To estimate /gg,. , a sample of organisms known to have just reached sexual maturity could be available and their arith- metic mean size can be used as an estimator. However, the sample needed to obtain such a design- based estimator (Smith, 1990) for wild populations might be too ex- pensive and would involve time-con- suming histological procedures. Fisheries biologists prefer to con- ceive size at first maturity as the average size at which 50% of the individuals are mature. With this conception, the estimator is not based on a sampling design but on a model of the relation between body size and the number of indi- viduals that are mature from a to- tal number at each of many size in- tervals. The variance of a design- based estimator is determined by sampling design (Thompson, 1992). The variance of a model-based esti- mator is not as easily obtained. A sample of published works in the fisheries literature provides a mea- sure of the frequency with which statistical uncertainty of the model- based /jQ,, is ignored (Table 1). In this work, we show three alterna- tive procedures; an analytical method derived from generalized linear models (McCullagh and Nelder, 1989), nonparametric boot- strap (Efron and Tibshirani, 1993), and a Monte Carlo algorithm devel- oped in our study. We show by simu- lation the behavior of the three methods for sample sizes of 500 to 10,000 individuals, concluding that they are similar in terms of bias, length, and shape of confidence inter- vals but that the Monte Carlo method outperforms the other two methods in percentage of times that the confi- dence interval contains the true pa- rameter, which remained close to the nominal 95% at all sample sizes. The problem In regression analysis, we are usu- ally interested in assigning confi- dence bounds to the response vari- able at specified levels of the pre- dictor variable. However, in matu- rity modeling the attention is turned to the converse problem of Roa et al.: Estimation of size at sexual maturity 571 Table 1 An example of published analyses on size at maturity in crustacean and fish popul ations. CW = carapace width; CL = carapace | length; TL = total length; m = males; f = females; s = gonadal maturity; " = morphometric maturity. Paper Species Fitting method '507, (inin) CI 959c CI estimation method Somerton Paralithodes Weighted nonlmear 102.8 (CL,m) Not reported Random partition of data (1980) camtschatica least-squares 101.9 (CL,f) Not repoited into subsets and computa- Chionoecetes bairdi 114.7 (CW. m) Not reported tion of var U^g,- ) among the N independent estimates of '507, Campbell Homarus americanus Nonlinear 108.1 (CL.fi and least-squares 92.5 (CL, f) — — Robinson 78.5 (CL,f) — — (1983) Somerton Paralithodes platypus Weighted nonlinear 80.6 (CL. f) 79.4 -82.6 Random partition of data and least-squares 96.3 (CL.f) 95.7 -96.9 into subsets and computa- Macintosh 93.7 (CL,f) 92.9 -94.5 tion of var (/^g,, ) among the (1983) 87.4 (CL.f) 86.4 -88.4 A'' independent estimates Campbell Cancer irroratus Nonlinear 62.0 (CW.m) and Eagles least-squares 49.0 (CL.f) — (19831 Somerton Lithodes aequispina Weighted nonlinear 97.7 (CL,f) Values not reported Bootstrap samples were and Otto least-squares 99.0 (CL.f) drawn from the original (1986) 110.7 (CL.f) 92.0 (CL. m) 107.0 (CL. m) 130.0 (CL. f) data set for obtaining in- dependent estimates of /gQ,., and then computing var(/jg,_ ) among them. Gaertner Geryon maritae Nonlinear 82.8 (CL, f) Confidence regions for the and Laloe least-squares parameters of a logistic (1986) function were computed. Comeau Chionoecetes opilio Nonlinear 34.2 (CW.m) and Conan least-squares (1992) Armstrong Lophius americanus Linear regression of 368.9 (TL.m) et al. (1982) (Pisces: Lophiiformesi Prop. Mature ( arcsine- square root trans- formed ) on Total Lengt h 458.3 (TL,f) Lovrich Paralomis granulosa Probit 50.2(CL, m)«5 and 60.6 (CL, f)E 58.3 -62.9 Not reported Vinuesa 57.0 (CL, ml" 53.9- -60.1 Not reported (1993) 66.5 (CL. f) 63.4- -69.5 Not reported Roa (1993a) Pleuroncodes monodon Maximum likelihood 27.2(CL, fi 24.2- -30.2 Ratio of parameter esti- mates confidence limits Gonzalez- Necora puber Maximum likelihood 54.8CW, m)e ^ Gurnaran 49.8"(CW, f)e — — and Freire 53.3 (CW.m)"" — — (1994) 52.3 (CW. f)™ — — setting a confidence interval for the size at which a fixed proportion of individuals in a population are sexually mature. That is, we need a procedure for estimating uncertainty in the predictor variable be- cause management decisions are framed in terms of body size, and hence the uncertainty in estimation 572 Fishery Bulletin 97(3), 1999 must be transferred to this variable. The first part of the problem is the selection of the maturity model. The available data consist of size (normally length) and maturity status, which will be assumed to take only two values: mature or immature. The predictor variable is continuous and the response variable is dichotomous. With such variables, model errors dis- tribute binomially. Welch and Foucher ( 1988) recog- nized this aspect of modeling maturity and showed an efficient procedure based on the principle of maxi- mum likelihood that takes advantage of the binomial nature of the errors. For dichotomous data modeled as a function of a continuous variable, the following simple logistic function is a consequence of the assumption of a lin- ear relationship between the logit link function and a single predictor variable (Shanubhogue and Gore, 1987; Hosmer and Lemeshow, 1989; McCullagh and Nelder, 1989): P(l): a 1 + e ii.*P,i :i) where P(l) = proportion mature at size /; and a, Pq, and p^ = asymptote, intercept, and slope pa- rameters, respectively (see also Eq. 3 ). The estimates of these parameters, given a data set, are chosen from the point at which the product of binomial mass functions of all data points (the likeli- hood of the data under the model) is a maximum, or equivalently when the negative of the log likelihood -na,pr,p,) = -^[(/!, )ln(P(/)) + (n, - /;, )ln(l- P(/))] (2) IS a minimum. these results, we may undertake the converse prob- lem of estimating size at fixed P^c maturity, which takes the form Pi 1 Ik (3) In Equation 3 it is assumed that the asymptote pa- rameter (a) from Equation 1 is fixed at 1. This as- sumption is justified on the basis of several published works on size at maturity, showing that all individu- als were mature above a given size during the repro- ductive season (Table 1). Furthermore, if P^^ and /3j are MLE of /i^ and /^^ and they are used to compute Ipr-^ from Equation 3, then /^^ is also MLE. We show below three procedures to perform this task and then test them by generating data from Monte Carlo simu- lation of the age-size structure and maturity progres- sion of individuals of a hypothetical population. Analytical estimation The logistic model in Equation 1 belongs to a class of generalized linear models studied by McCullagh and Nelder (1989). These authors consider the problem of building approximate confidence intervals for the level of the predictor variable that gives rise to a fixed proportion in the response variable. They suggest the use of Fieller's (1944) theorem, according to which the linear combination P,, + p,lp.-,-g(P^) = Q, (4) where Ip,^ = the value of the predictor variable for a fixed proportion; and g(P^y)=ln(P^^/(l-PQ)) (the logit link function) is approximately normal with mean zero and analytical variance given by where /i - thenumber of mature individuals; and n = sample size at /; P(lJ = Eq. 1; and where a constant term that does not affect the esti- mation is omitted. Given the nonlinear nature of normal equations, the minimum is found by an iteration algorithm. The parameters estimated by minimizing Equation 2 are maximum likelihood estimates (MLE). In practical situations, the logistic model may be modified from its original form to allow more biological reality (Welch and Foucher, 1988). The result from fitting the model (Eq. 1) to the data by using the objective in Equation 2, is a vector of parameter estimates and a covariance matrix, which represents the uncertainty associated to them. With v'^ilpr.) = vaT{Po) + 2lp,; cov{Po,Pi) + Iprr, var(^i). (5) The lOO(l-a)'^ confidence interval is the set of val- ues defined by Pi where z^^/., - a quantile of the normal distribution. Other link functions like probit, common in the field of toxicology (Finney, 1977), are not investigated in this paper. Bootstrap estimation Bootstrap is not a uniquely defined concept (Efron and Tibshirani, 1993). This means that bootstrap I Roa et al : Estimation of size at sexual maturity 573 samples may be obtained by conceptually different resampling procedures. In the context of logistic re- gression, it is possible to resample the observational pair (/,,/!,), or the semiobservational pair (Z^, P(l)+£^), with fj as a realization from the residual distribu- tion of the logistic model. To be valid, this second resampling unit needs the assumption of indepen- dence between e^ and l^. As stated by Efron and Tibshirani (1993), this is a strong assumption that can fail even when the model P(l) is correct. These authors remark that bootstrapping the obsei-vational pair is less sensitive to assumptions than bootstrap- ping residuals. Therefore, in our work an observa- tion to be resampled with replacement is defined as the pair (length, maturity status). For each and all bootstrap samples, a resampled frequency distribu- tion for Ipr. is obtained by fitting the maturity model in Equation 1 with the objective function in Equa- tion 2 and by computing Ip, with Equation 3. The confidence interval is obtained by application of the bias-corrected and accelerated (BCa) method, recom- mended by Efron and Tibshirani (1993). Monte Carlo estimation In Monte Carlo resampling, a model is assumed for the distribution of the estimator and then data are generated computationally to assess the amount of variation (Manly, 1997). In our case, we consider a Monte Carlo resampling of maturity parameters from the modeled joint probability distribution of the es- timates /^Q and /Jj for computing / from Equa- tion 3. In contrast to the bootstrap approach, the implementation of this approach needs only one fit- ting of the logistic maturity model and then uses the asymptotic distribution of estimated parameters of the model to generate the probability distribution of the derived statistic Ip... These parameter estimates, ^Q and /}j, distribute asymptotically bivariate nor- mal, with mean vector equal to the population pa- rameters and variance given by their covariance matrix (for nonlinear least-squares: Johansen, 1984; for maximum-likelihood estimates: Chambers, 1977 ). The bivariate normal distribution of p^ and /3j has a strong covariance component, which is the same as to say that these estimates are highly correlated. This also means that much of the variance in one estimate is given by the variance in the other one. Ignoring such correlation would lead to an overesti- mation of the variance oi Ip,.. In a Monte Carlo set- ting, the correlation between parameter estimates may be considered in the computation by making the resampling of one estimate conditional on the resampling of the other one. In this work we develop such a technique using the theory of least-squares estimates of two linearly related normal variates (Draper and Smith, 1981). This approach is justified by the asymptotic nature of standard errors. If /Jq and j8j and are two normal random variables that are linearly related, then we may write the linear equation Pi=bo+b^Po- (7) This equation may be reversed by writing /?j, as a linear function of p^ because both are random vari- ables. It can be shown that (Draper and Smith, 1981) bi Ml \^.] h\ (8) where /■ is^ the estimated linear correlation coefficient between /3q and P^, and S/jq and S p^ are the respec- tive standard errors. Furthermore, from Equation 7 bo=p,-b,p,. (9) Therefore, the high correlation coefficient between both maturity parameters can be accounted for by free sampling from the marginal distribution of one parameter estimate (for example, /Jq) in each Monte Carlo trial and by computing the other by using Pi,=Pi-t A,. A Pi + r AA ^. A, Ao ^30+^^, Ao.Ai ^0,,-A, A, Po.j , (10) which is obtained by replacing Equations 8 and 9 in Equation 7. For each trial (indexed byj), a /?y value is selected from the normal probability distribution defined by its estimate and standard error, and then the mean P^ value is computed by using Equa- tion 10. The variance of the estimate is the p^ variance due to the linear relationship with /^^ plus a residual variance not explained by the relationship. The vari- ance due to the relationship is directly transferred from ^Q to /}j through the Monte Carlo resampling of /3q and its mapping onto ^j by using Equation 10. The residual variance must be added in each trial with ^p^, residual Ao.Ai (11) 574 Fishery Bulletin 97(3), 1999 where r~ ■ = the proportion of variance due to the Po.P hnear relationship. Note in Equation 10 that when r=0, the mean of the ^j for all^ would just be the /?j estimate, which means that /^^ and /3, values are independently se- lected in each trial; note also in Equation 1 1 that the resampling variance of the estimate would be its to- tal variance. On the other hand, when r= 1 1 1 , Equa- ./tion 11 shows that the resampling variance of /3j would be totally due to the mapping of p^ ^ onto p.^ ^, which is expected when the linear relationship be- tween two variables is deterministic. In this case, the algorithm presented here would only perform one Monte Carlo simulation, that on /J,,. Therefore, the algorithm is flexible enough to cover the whole range of correlation between both parameter estimates. A confidence interval for Ip,- may be obtained by the percentile method (Casella and Berger, 1990; Efron and Tibshirani, 1993), for which two computa- tional alternatives are available. If the resampling through the bivariate normal distribution is un- bounded, then the 100( l-a)% confidence interval is obtained by ordering the lp<,j from smallest to larg- est, and taking as bounds the values at positions Nmc(oc/2) and NMc(l-(oil2), where N}^c is number of Monte Carlo trials. If the resampling through the bivariate normal distribution is bounded, with bounds al2 and l-aJ2, then the 100(l-a)% confi- dence interval limits are obtained as the first and last quantiles when ordering the //>.,; ^ from smallest to largest. Monte Carlo simulation To test the performance of the three procedures in estimating Ip,. for different sample sizes, we carried out a simulation analysis of a model population with known size-at-age structure, maturity-at-size, and mortality parameters ( Table 2 ). We explored only the behavior of the methods for median (50%) size at maturity (/^^,, ). Performance was evaluated by us- ing four criteria. First, as the proportion of times that confidence intervals did contain the true (and known ) parameter (Ir^Qc,^ ), which we call success: success = 1 - failure number\(true -luwer)iupper - true) < 0} number of iterations (12) Our second criterion was bias, evaluated as the av- erage, over trials, of the sufficient statistic: Table 2 Parameter estimates used in the model to generate simu- lated data. Growth and maturity functions given by Roa ( 1993a) for female squat lobsters iPleuroncodes monodon ). Natural mortality rate (M) given by Roa (1993b) for the same species. Parameter Value Size-at-age ff2 4 Growth L k (/yr) 'n(/yrl 44.55 0.179 -0.51 Maturity a A, A, '.w, 'mm) 1 13.648 -0.502 27.2 Natural mortality M (/yr) 0.6 bias resampled median (13) true which is 1 for an unbiased estimator. The third crite- rion was the length of confidence intervals: length = upper - lower. (14) and the fourth and final criterion was the shape of the interval (Efron and Tibshirani, 1993): shape upper - median median -lower (15) which measures asymmetry around the median. In all four measures of performance, "upper" and "lower" refer to the bounds of the confidence interval, "me- dian" is the median Ipc^, and "true" refers to the true value. The deterministic and stochastic features of our simulation were chosen for a population with features like those previously reported for the squat lobster iPleuroncodes monodon) from the continen- tal shelf off central Chile (Roa, 1993a, 1993b). To accomplish this task, we implemented the fol- lowing three-step algorithm, which we called MATSIMVL: step 1, generation of A^,^,,^=5000 random samples of maturity-at-size data of sample size N , = 500, 1000, 3000, 5000, and 10,000 individu- sample als (Eqs. 16-19); step 2, estimation of the parameter vector and covariance matrix for each one of these samples (Eqs. 1 and 2); and step 3, running each of the three methods to obtain the 2.5%, 50%., and 97.5%' Roa et al,: Estimation of size at sexual maturity 575 percentiles of /gy,_ (Eqs. 3-11) In bootstrap, for each sample size and each of the 5000 trials, we obtained N 1^^^=5000 bootstrap samples. With the Monte Carlo method, we resampled parameter estimate values from unbounded normal distributions with Nj^^-.^ 5000. For completeness, 7V^^p^,^,^= 1 . In step 1, the deterministic size structure of the population was conceived as a mixture of normal probability distributions, each normal distribution corresponding to an age class. The proportion of in- dividuals at each size interval was characterized by the following expression ran a is the random number of mature individuals V P"< - 2I 0 j P"i 40 9 SI Ay- /=o /=o ■v2;rCT P 2I a J (16) where the sum is over 10 age classes (0 to 9) and 41 size classes (0 to 40), and where f.i^ is determined by a growth equation ^, ^R.(l-e-'-) (17) with known parameters (Table 2). Variance of size- at-age (a'~) is known and constant through age (Table 2), and the proportion of individuals at age (P^^) is given by a simple exponential mortality model -Ml -Ml (18) t=0 where the mortality rate (M) is known and constant through age (Table 2). Random variability came from two sources. First, samples of the specified sizes were drawn, for each trial, from a uniform probability distribution and compared with the cumulative distribution of Equa- tion 16, accumulating the scores in the respective size intervals. This computation yielded a sample of relative size frequencies p^^^. Next, we introduced the second source of uncertainty by assessing the matu- rity status (mature or immature) of individuals be- longing to each size class. This random assignment of maturity status came from resampling the bino- mial probability distribution P(n = n rand U.rand ^rand / pa )"-'(l-P(/)'""°'"'-""'"''),(19) where Pf/i was computed from the logistic model (Eq. 1) with known maturity parameters (Table 2) and out of «, I. rand sample ^ nl individuals in the size in- terval /. In this way, step 1 was completed by ran- domly assigning two properties to each data indi- vidual: a size (continuous variable) and a maturity status (dichotomous variable). With these data, step 2 was completed by using a nonlinear parameteriza- tion of the logistic model (Eqs. 1 and 2) for obtaining estimates of /3q and (i^, and their covariance matrix, by means of the SIMPLEX algorithm (Press et al., 1992). Having this information in hand, step 3 was completed by obtaining 2.5%, 50%, and 97.5% per- centiles by each of the three methods. We pro- grammed the MATSIMVL algorithm using Microsoft FORTRAN for PowerStation 4.0 (Microsoft Corp., 1995). In the case of the Monte Carlo algorithm, we also investigated the effect of the natural mortality pa- rameter, by varying its level in simulation at M=0.2, M=0.4, M=0.6, and M=0.8, for sample sizes of N„,„^,,^=1000 and 5000 individuals. N,^,,. and N^^ were both kept at 5000. Finally, we introduce real data to show two appli- cations of the Monte Carlo method developed here. First, we estimate Ip,^ (^^,^.=5000) for a single popu- lation of the galatheid decapod Pleitroncodes monodon. In this application, we estimate size con- fidence intervals for percentages of maturity from 10% to 90% at steps of 10% . In this way a confidence interval for the whole maturity curve is outlined. Second, we compare samples of female anchovy Engraulis ringcns from two localities 3^ of latitude apart (N^,(^.=5000) to test the null hypothesis of equal /jQ,, between them. Results The simulation analysis with MATSIMVL yielded size-at-age and maturity-at-size data with the ap- propriate behavior as Ng^„„i^ increased: size-fre- quency distributions became smoother and maturity data more closely followed a logistic curve, as shown by one example output of MATSIMVL data-simulat- ing routines (Fig. 1). A summary of the simulation results is presented in Fig. 2. It shows that, under the simulation condi- tions, the Monte Carlo method outperformed the bootstrap and the Fieller methods in proportion of success at all sample sizes and that it remained very close to the nominal 95%; the bootstrap method suc- ceeded 94% or less at all sample sizes, whereas the Fieller method was unstable between sample sizes of 500 to 5000, with a minimum of 93% success at 3000 (Fig. 2A). All three methods showed negligible 576 Fishery Bulletin 97(3), 1999 o D. O 1.0 0.5 00 1.0 0.5 ao 1 0 c 0 5 - o 0.0 10 0,5 ■ ao 10 0 5 0.0 B D H M"'M***'<' 0 5 10 15 20 25 30 35 40 45 Size (mm) Figure 1 Example output of simulated data on size frequency and maturity-at-size generated by one trial of the MATSIMVL algorithm, for sample sizes of 500 (A and B ), 1000 (C and Dl. 3000 (E and Fi. .5000 (G and Hi. and 10,000 (I and J) individuals. bias at low sample sizes and converged neatly to null bias at large sample sizes (Fig. 2C). Likewise, the three methods behaved exactly the same in length of confidence interval, decaying exponentially as sample size increased (Fig. 2C). Finally, both resampling methods yielded asymmetrical (right- tailed) confidence intervals, converging to the same shape value (ca. 1.1) as sample size increased (by definition, Fieller's method yields a symmetrical con- fidence interval with shapes 1). Percentage success and bias of the Monte Carlo method under our model for data generation were fairly insensitive to changes in natural mortality M (Fig. 3) for sample sizes of 1000 and 5000 individu- als. Percentage success remained close to the nomi- nal 95'7( and bias was negligible. The length of the confidence interval and asymmetry, however, in- creased with increasing mortality, showing that es- timation variance was directly proportional to natu- ral mortality rate. Results of the Monte Carlo algorithm with real data on female squat lobster are shown in Figure 4. The Monte Carlo confidence interval for /r,,,,; was fairly narrow (25.86 to 28.51 mm carapace length), Roa et a\ : Estimation of size at sexual maturity 577 o C/5 I 000 -, 0975 0950 0.925 0900 1 .0050 T 1.0025 2000 4000 6000 — I 1 8000 10000 2000 4000 6000 8000 10000 ■i 1.0000 CD 0 9975 - 0 9950 - I 1.5 - 14 - 13 - « 12- on 11 - 10 - 09 ■ Sample size B 2000 4000 6000 8000 10000 2000 4000 6000 8000 1 10000 Figure 2 Summary of results from the application of the three methods to simulated data. Open squares = Fieller's ana- lytical method; open circles: bootstrap; closed circles = Monte Carlo. reflecting the effect of large sample size 'A^,q„, i^ = 4458). The Monte Carlo median /-^.^ (27.19 mm cara- pace length) coincided with the MLE of /r,iy;=-MLE (^„)(MLE(/3j ). The amplitude of the confidence inter- val for Ip, showed an increasing trend towards ex- treme values of P (Fig. 4), a reflection of the alge- braic structure of Equation 3. Results of the second application with two samples of female anchovies are shown on Figure 5. Although different in their lengths, probably due to different sample sizes, con- fidence intervals from the two samples overlapped and have the same upper limit. This result provides sup- port to the hvpothesis of equal maturity schedules be- tween female anchovies from the two localities. Discussion The logistic model is universally used as a math- ematical description of the relation between body size and sexual maturity. To model residuals, however, two different approaches emerge: to consider them normally or binomially distributed, and closely re- lated to this, to use the data as proportions or as counts. In the first (as far as we know) formal treat- ment of the problem. Leslie et al. (1945) used the data as proportions, transformed to probit scores, and assumed the normal distribution. Current research- ers have not employed probit transformations (but see Lovrich and Vinuesa, 1993) but have continued using data as proportions and the normal distribu- tion for residuals (Table 1). In this work however, and following the arguments by Welch and Foucher (1988), we emphasize the need to estimate the ma- turity model using the data as counts and therefore to consider residuals as binomially distributed. Un- der this approach, the standard procedure for fitting the maturity model is logistic regression (Hosmer and Lemeshow, 1989). When the model has been fitted, the problem of setting a confidence interval for the level of the pre- dictor variable (size) that gives rise to a fixed pro- portion of maturity is not trivial. We have explored here three approaches: one analytical method based 578 Fishery Bulletin 97(3), 1999 3 C/5 1.000 0,975 - 0.950 - 0925 - ■s H 00 1 - CQ 1,0050 T 1 0025 1.0000 - 0 9975 - B 0,900 ^ 1 1 1 1 1 1 1 1 0 9950 H 1 1 1 1 1 1 1 ; 0 1 0,2 0,3 0.4 0 5 0.6 0 7 0 8 0 9 0 1 0 2 0 3 0.4 0.5 0.6 0 7 0 8 0 9 4 -r — I 1 1 1 1 1 1 1 0.1 0.2 0.3 0.4 0.5 0,6 0.7 08 09 1 1~ 0 1 0.2 0 3 0 4 0.5 0.6 0.7 0.8 0 9 Natural Mortality Figure 3 Effect on natural mortality rate on the performance of the Monte Carlo method. Open circles =1000 indi- viduals; closed circles = 5000 individuals. A^,„„,p,,= 5000, N^i^= 5000. on Fieller's (1944) theorem and the logit hnk func- tion, and two computationally-intensive methods based on nonparametric bootstrap of the observa- tional pair (/,,^,l and Monte Carlo resampling of pa- rameter estimates from the logistic model. Although this is not an exhaustive set of methods (for example, we did not explore likelihood profiles), they repre- sent a set of conceptually different alternatives to be tested against the model we used to generate simu- lated data. In particular, both resampling methods are especially useful to obtain the distribution of a function of estimated parameters, such as in Equa- tion 3, mainly because of their mathematical sim- plicity, which comes at the expense of extensive com- putation. Our results indicate that the three meth- ods to estimate size at P9c maturity perform almost equally well in terms of bias, length, and shape of the confidence interval, but that Monte Carlo per- formed better in containing the true parameter within its confidence bounds with the nominal 95% rate. This greater accuracy is accompanied by Nf^^^^-1 times less computation than bootstrap. Bootstrap single assumption was that all observa- tions from any given sample have the same prob- ability to appear in a new sample. In contrast, the Monte Carlo method assumed a bivariate normal distribution of parameter estimates of the maturity model. Having simpler assumptions, it is unclear why the bootstrap method failed more than the nominal 5% of the times at low sample sizes. One reasonable explanation is that for every sample, there would be A^^^^^^, bootstrap samples, and therefore A^^^^^j, numeri- cal solutions to the normal equations under the bi- nomial likelihood model. We used here 5000 boot- strap samples and the SIMPLEX algorithm (Press et al., 1992). Small errors in the numerical algorithm coupled with a minimum bias, may add to the nomi- nal 57( , accounting for the I7r to 2% increase in fail- ure rate. In contrast, the Monte Carlo algorithm re- quires a single numerical solution so that it does not accumulate numerical errors. On the other hand, Fieller's analytical method requires a more compli- cated set of assumptions than the Monte Carlo ap- proach. Fieller's method requires normality of a lin- Roa et a\: Estimation of size at sexual maturity 579 ^y^o o" / ° 0.8 - / / ° o / i 0.6- c / ■ Proportion o 1 ■ / / ' / 0.2 - / ' / * / ° ^ T-^^ 15 20 25 30 35 40 45 50 Carapace length (mm) Figure 4 Monte Carlo confidence inter\'als for P'> maturity for female squat lob- sters, Pleuroncodt's monodon 1 A'^^,,^^,^ =4458, jV,„.=5000 >. Open circles = raw- data; continuous line = fitted model; closed squares = Monte Carlo confi- dence bounds for length at P''< maturity iP=0. 0-1.0 in steps of 0.1). ear combination of parameter estimates and therefore assumes a symmetric in- terval estimate. Both bootstrap and Monte Carlo results indicate that the interval estimate can be quite asymmet- ric. This may account for the better per- formance of Monte Carlo, as compared with Fieller's, in proportion of success. These remarks, along with the facts that Monte Carlo's percentage success and bias are not affected by a natural mortality rate varying between 0.2 and 0.8, and that it is very fast on every com- puter platform, allows us to recommend the use of the Monte Carlo method to estimate size at P'/f maturity. Acknowledgments We would like to thank Bryan F. J. Manly for reviewing an earlier draft of this work and three anonymous review- ers who made several comments and criticisms which greatly improved the manuscript. This work was funded by FONDECYT grant 1950090 to R.R. 9 10 11 12 Standard length (cm) Figure 5 Monte Carlo confidence inten-als for length at 50<7f maturity of female anchovies {Engrauhs r(;igt>ns) sampled from two localities off central Chile. Open circles and dotted line = Talcahuano (36 45'S) (^ sampk''^^'^- A'„^=5000l; Closed circles and solid line liV San Antonio (33°35'S) =585.iV,„,=5000); Closed squares = 95<7t confidence bounds for/jg,,. 580 Fishery Bulletin 97(3), 1999 Literature cited Armstrong, M. P., J. A. Musick, and J. A. Convocoresses. 1992. Age, growth, and reproduction of the goosefish Lophius americanus (Pisces: Lophiiformes). Fish. Bull. 90:217-230. Campbell, A., and M. D. Eagles. 1983. Size at maturity and fecundity of rock crabs, Cancer irroratus, from the Bay of Fundy and southwestern Nova Scotia. Fish. Bull. 81(21:357-362. Campbell, A., and D. G. Robinson. 1983. Reproductive potential of three American lobster iHoinarus americanus) stocks in the Canadian maritimes. Can. J. Fish. Aquat. Sci. 40:1958-1967. Casella, G., and R. L. Berger. 1990. Statistical inference. Brooks and Cole Publ. Co., Duxbury, California. 650 p. Chambers, J. M. 1977. Computational methods for data analysis. John Wiley and Sons, New York, NY, 268 p. Comeau, M., and G. Y. Conan. 1992. Morphometry and gonad maturity of male snow crab. Chionoecetes opilio. Can. J. Fish. Aquat. Sci., 49:2460- 2468. Draper, N. R.. and H. Smith. 1981. Applied regression analysis. John Wiley and Sons, New York. NY, 736 p. Efron, B., and R. J. Tibshirani. 1993. An introduction to the bootstrap. Chapman & Hall, New York. NY, 436 p. Fieller, E. C. 1944. A fundamental formula in the statistics of biological assay, and some applications. Quart. J. Pharm. 17:117- 123. Finney, D. J. 1977. Probit analysis, 3rd ed. Cambridge Univ. Press, New York, NY, 333 p." Gaertner, D., and F. Laloe. 1986. Etude biometrique de la taille a premiere maturite sexuelle de Geryon mantae Manning et Holthuis, 1981 du Senegal. Oceanol. Acta 9:479-487. Gonzalez-Gurriaran, E., and J. Freire. 1994. Sexual maturity in the velvet swimming crab AAecora puher (Brachyura, Portunidael: morphometric and repro- ductive analyses. ICES J. Mar. Sci., 51:133-145. Hosmer, D. W., and S. Lemeshow. 1989. Applied logistic regression John Wiley and Sons, New York, NY'. 328 p. Johansen, S. 1984. Functional relations, random coefficients, and non- linear regression with application to kinetic data. Springer- Verlag, New York, NY, 126 p. Leslie, P. H., J. S. Perry, and J. S. Watson. 1945. The determination of the median body-weight at which female rats reach maturity. Proc. Zool. Soc. Lond. 115:473-488. Lovrich, G. A., and J. H. Vinuesa. 1993. Reproductive biology of the false southern king crab iParalomis granulofia. Lithodidae) in the Beagle Channel, Argentina. Fish. Bull. 91:664-675. Manly, B. F. J. 1997. Randomization, bootstrap and Monte Carlo methods in biology. 2"'' ed. Chapman & Hall, London, 399 p. Microsoft Corp. 1995. FORTRAN PowerStation programmer's guide. Printed in the USA, 752 p. McCullagh, P., and J. A. Nelder. 1989. Generalized linear models. Chapman & Hall, Lon- don. 511 p. Press, W. H., S. A. Teukolsky, W. T. Vettering, and B. P. Flannery. 1992. Numerical recipes in FORTRAN, 2"'' ed. Cambridge LIniv. Press, Cambridge, 963 p. Roa, R. 1993a. Annual growth and maturity function of the squat lobster Pleuroncodes monodon in central Chile. Mar Ecol. Prog. Sen 97:157-166. 1993b. Analisis metodologico pesqueria de langostino Colorado. Tech. Rep., Instituto de Fomento Pesquero I IFOP), Chile. 86 p. (Available from .Subsecretaria de Pesca, Bellavista 168, piso 19, Valparaiso, Chile.) Shanubhogue, A., and P. A. Gore. 1987. Losing logistic regression in ecology. Curr Sci. 56; 933-936. Smith, S. J. 1990. Use of statistical models for the estimation of abun- dance from groundfish trawl survey data. Can. J. Fish. Aquat. -Sci. 47:894-903. Somerton, D. A. 1980. A computer technique for estimating the size of sexual maturity in crabs. Can. J. Fish. Aquat. Sci., 37:1488- 1494. Somerton, D. A., and R. A. Macintosh. 1983. The size at sexual maturity of blue king crab, Paralithodex platypus, in Alaska. Fish. Bull. 81(3):621- 628. Somerton, D. A., and R. S. Otto. 1986. Distribution and reproductive biology of the golden king crab, Lithodes aequispina, in the eastern Bering Sea. Fish. Bull. 84(3):571-584. Thompson, S. K. 1992. Sampling. John Wiley & Sons, New York, NY, 343 p. Welch, D. W., and R. P. Foucher. 1988. A maximum likelihood methodology for estimating length-at-maturity with application to Pacific cod iGadus macrocephalus) population dynamics. Can. J. Fish. Aquat. Sci. 45:333-343. 581 Abstract.— Patterns of growth and mortality were examined for post- settlement red drum, Sciaenops ocel- latits. inhabiting seagrass meadows in the Aransas Estuary, Texas. Age and growth rates of larvae and early juve- niles were estimated in 1994 and 1995 by using daily increments in otoliths. Otolith-derived estimates of age indi- cated that individuals spend approxi- mately 20 d in the pelagic environment before entering demersal habitats (i.e. before settlement). Instantaneous growth coefficients (g) of red drum ranged from 0.049 (4.8'7f/d) in 1994 to O.OSKS.C^f/diin 1995. Site-specific dif- ferences in growth were also examined and a significant site effect was de- tected in 1994; however, no site effect was observed in 1995. Interannual and cohort-specific ( 10-d cohorts ) mortality rates were estimated from declines in log^, abundance (abundance-at-age plots), and results indicated that mor- tality during the early postsettlement period was substantial. Instantaneous mortality coefficients (Z) were similar between years (0.134 IVI.S^Ud] in 1994; 0. 139 1 13.0'7f /dl in 1995 ), and no signifi- cant interannual effect was observed. Conversely, cohort-specific mortality rates ranged widely (0.106-0. 265 1 10.1- 23.3'-^/d] ) and losses were lowest for midseason cohorts. Recruitment poten- tial {G:Z ratio) was highest for mid- season cohorts (1.30-1.56) and lowest for early and late-season cohorts (<1). Although G:Z ratios varied over spatial and temporal scales, ratios were >1 in 1994 and 1995, suggesting that both year classes experienced favorable nursery conditions. Spatial and temporal variability in growth, mortality, and recruitment potential of postsettlement red drum, Sciaenops ocellatus, in a subtropical estuary* Jay R. Rooker Scott A. Holt G. Joan Holt Lee A. Fuiman Marine Science Institute The University of Texas at Austin 750 Channelview Drive Port Aransas, Texas 78373 Present address (for J R Rooker) Department of Marine Biology Texas A&M University 5007 Avenue U Galveston, Texas 77551 E-mail address (for J R Rool^er) rookeriiatamug lamu edu Manuscript accepted 27 August 1998. Fish. Bull. 97:581-590 (1999). Survival during the lai-val and ju- venile stages of marine fishes is highly variable and plays a critical role in determining recruitment potential (Gushing, 1975; Houde, 1987). Early life survival is a func- tion of both gi-owth and mortality, and it is clear that these mecha- nisms act in concert to determine an individual's probability of sur- vival (Houde, 1996). Individuals experiencing rapid growth will spend less time in vulnerable size ranges (reduced stage duration) and achieve a larger body size at a given age, thus enhancing their ability to detect and escape predators (Bailey and Houde, 1989; Fuiman and Magurran, 1994), Consequently, recruitment success or failure is closely linked to variation in growth and mortality during the early life stage. Much of our current understand- ing of early life growth and mortal- ity is based on information derived from the examination of otolith mi- crostructures. The daily deposition of growth increments on otoliths has been demonstrated for many ma- rine teleosts and provides a means of estimating age, growth rate, and hatching date (see reviews by Campana and Neilson, 1985; Jones, 1986). Moreover, knowledge of a population's age structure can be combined with density data (abun- dance-at-age relationships) to esti- mate mortality rates. Thus, funda- mental demographic parameters can be obtained through otolith analysis, and these data are essen- tial for determining the causes and consequences of differential survival. Red drum {Sciaenops ocellatus) inhabit subtropical and temperate waters in the Western Atlantic and support important commercial and recreational fisheries throughout the coastal waters of the Gulf of Mexico (Swingle, 1990; NOAA, 1991 ). Red drum spawn in the early fall in offshore waters and areas near tidal inlets, and tidal currents transport larvae through passes and into estuarine nursery habitats (Holt et al., 1989; Comyns et al,, 1991). Individuals settle into sea- * Contribution 1060 of Marine Science Institute, The University of Texas at Aus- tin, Port Aransas, TX 78373. 582 Fishery Bulletin 97(3), 1999 grass and marsh-edge habitats following a fairly short pelagic interval and remain in these habitats during the juvenile stage (Baltz et al., 1993; Rooker and Holt, 1997; Rooker et al., 1998b). Although much effort has been directed to- ward the management of red drum stocks, re- search on growth and survival during early life has received limited attention. As a result, ihe aim of this study was to examine spatial and temporal patterns of growth and mortal- ity during the postsettlement period. Specific objectives of this research were to estimate annual variation in growth and mortality, to estimate cohort- and site-specific mortality rates, and to determine the recruitment po- tential of different cohorts. Materials and methods Field collections Red drum were collected from two shoal grass [Halodule ivrightii) meadows in the Aransas Estuary, Texas (Aransas Bay 1 and 2 [ABl, AB2]; Fig. 1), during the annual settlement period (August-December). Collections were taken over a two-year period (1994-951 with an epibenthic sled measuring 0.75(w)x0.5(h) m, equipped with a 505-)am mesh conical plankton net. Triplicate 20-m sled tows were taken every week at each site in 1994. A more comprehensive sampling strategy was em- ployed in 1995 in order to evaluate cohort-specific variation in natural mortality. Ten 20-m sled tows were taken every 3-4 d at each site during the en- tire 1995 settlement season. Red drum larvae and juveniles were preserved in 70% ethanol immediately after capture. Environmental data collected at each site included depth, salinity, and water temperature. Laboratory procedures Removal and processing of otoliths (lapilli) followed the procedures described by Rooker and Holt ( 1997). Otolith radius, increment count, and increment widths were measured on a straight line from the core to the posterior edge using an image analysis system (Optimas, Bioscan). Because otolith growth is allometric (Campana and Neil'son, 1985), all mea- surements were taken in the same field. Daily deposi- tion of increments was validated in our laboratory by chemically marking the otoliths of known-age red drum with alizarin complexone (S.A. Holt, unpubl. data). irsA- 27 'SO' 47 06' Figure 1 Location of shoal grass (Halodule wrightii) sampling stations (ABl and AB2) in the Aransas Estuary. Texas. Age was determined by enumerating growth in- crements from the core to the margin of the otolith. Inner increments (proximal to the core) on lapilli were often difficult to enumerate. Thus, in order to accurately determine the age of wild red drum, a re- lationship between age and otolith radius from labo- ratory-reared red drum was developed and used to predict the number of growth increments at various distances from the core ( Rooker and Holt, 1997 1. Age was determined by adding the predicted age from the unreadable section (correction factor from age- radius relationship) to the increment count (number of increments from first identifiable increment to edge of the lapillus). Correction factors generally accounted for less than 25% of the actual age esti- mate, and the relative size of the correction did not differ between years. Both the left and right lapilli were examined and estimates of age and growth were derived by averaging readings from both otoliths. On average, age differences between otoliths were small (CV=3.6%: range 0-14%). For some individuals Rooker et al.: Variability in growth, mortality, and recruitment of Sciaenops ocellatus 583 (18%), only one of the lapilli was available owing to loss, breakage, or staining. Growth rates were determined by using otolith- derived estimates of age (size-at-age plots) in 1994 (n=249) and 1995 (n = 140). Daily instantaneous growth coefficients were calculated from an exponen- tial model described as where L^ = length (mm SL) at time t; Ly = the estimated length at hatching; g = the instantaneous growth coefficient (/d); and t - theotoKth-derivedage(daysafterhatching). Ages of red drum not assessed with otolith-based techniques ( 1994 [« = 1,057], 1995 [n=8004] ) were es- timated by using age-length relationships. Weight-specific instantaneous growth coefficients (G) were calculated by using the equation W^ = W^e^', where W^ = the wet weight (mg) at time ^ Wg = the estimated weight at hatching; G = the weight-specific instantaneous growth coefficient (/d); and t = the otolith-derived age. Red drum lengths ( mm ) were converted to wet weight (mg) by using a polynomial equation (third order) based on measurements from laboratory-reared red drum ( range; 6-30 mm ): W = -7.745 -t- 2. 122L - 0.205L- -t- 0.024L'^ (n=200, r"=0.97), where W and L represent wet weight and standard length, respectively. Mortality rates were estimated from regressions of the decline in log, -transformed abundance on age. Although interannual variation in mortality was examined by pooling individuals from each year (all cohorts combined), cohort-specific rates were based on regression plots of log^.-transformed abundance of individuals from 10-d cohorts determined from hatching-date analysis. Hatch dates of individual red drum were determined by subtracting the otolith- derived age from the date of collection. Hatching dates were then used to separate individuals into specific cohorts, defined as individuals hatched within a 10-d period. Each cohort was designated by a letter (A to F): 1-10 Sep (A); 11-20 Sep (B); 21-30 Sep (C); 1-10 Oct (D); 11-20 Oct (E); and 21-30 Oct (F). Densities for the first and last cohort (A, F) were low and thus mortality regressions were not fitted to these data. Daily instantaneous mortality rates were calculated fi-om the exponential model of decline: N, = N^e-^', where A^^ = abundance at time t; Nq = the estimated abundance at hatching; Z = the instantaneous mortality coefficient (/d); and t = the otolith-derived age. Owing to incomplete capture (i.e. ascending limb of catch curve) of small red drum (<25 d, <8 mm), these individuals were not included in mortality re- gressions. Also, size-based gear avoidance was sus- pected, which placed constraints on the upper age (or size) of individuals used in regression analysis. In preliminary trials, the capture efficiency of the epibenthic sled was compared with a bag seine (seine dimensions: 7 m length x 1 m height; mesh size: 3 mm) and size-specific differences in capture effi- ciency (individuals/m''^ of habitat sampled ) between the two gears were not detected for sciaenids <25 mm (Rooker, 1997). However, density estimates of sciae- nids >25 mm were higher with the seine. This sug- gested that individuals >25 mm are capable of avoid- ing the sled and that densities calculated for these individuals would be underestimated. Consequently, only red drum between 8 and 20 mm (25-40 d ) were used to estimate mortality rates. Our assessment of early life mortality was based on two assumptions: 1 ) individuals entering seagrass sites remained in these habitats (i.e. settle and stay) during the time period when mortality rates were estimated, and 2) immigration and settlement (i.e. late settlers) to the designated study sites from other locations or habitats, or both, was negligible. Al- though postsettlement movement (emigration or immigration, or both) can be important for certain species and failure to account for such movements can severely bias mortality estimates (Frederick, 1997), our assumptions are reasonable because red drum appear to settle and stay in seagrass meadows at least through the early juvenile stage (Rooker et al., 1998b). Furthermore, no salient increases or de- creases in density were observed in length-frequency profiles which would be expected in the presence of postsettlement emigi'ation or immigration activity. Since total mortality (Z) is the sum of both natural (M) and fishing mortality (F), it is important to ac- count for the fishing mortality iF) caused by our sam- pling (Ricker, 1975). Most studies evaluating early life mortality do not estimate F (^sampling mortal- ity ) and assume F to be inconsequential. In this study, seagrass meadows of limited size (ca. 25,000 m-) were sampled repeatedly and, as a result, the effect of F was considered potentially significant. Therefore, mortality regressions were run on abundance data 584 Fishery Bulletin 97(3), 1999 adjusted to compensate for sampling losses (F). Abun- dance at each sampling time was adjusted by sub- tracting the number of individuals collected ( removed from population) with the epibenthic sled from the overall abundance at the site (site abundance = area of site/area sampled xnumber of red drum collected). Regressions of log^-adjusted abundance on age were then run and mortality coefficients derived in this manner were basically estimates of M since the ef- fect of F was removed. Consequently, F was deter- mined by subtracting adjusted mortality rate (M) from Z. Because F values were relatively small, 2.3 to 6.9% of Z, only estimates of Z are presented. The relative recruitment potential of individual cohorts was assessed by examining the ratio of weight-specific growth (G) to mortality (Z). The ra- tio is commonly used as an index of stage-specific survival because it incorporates both growth and mortality (Werner and Gilliam, 1984; Houde, 1996). Because cohorts with G:Z ratios >1 gain biomass, the probability of survival (recruitment potential) for these individuals is assumed to be high. Data analysis Analysis of covariance (ANCOVA) was used to test for intra- and interannual differences in growth and mortality (covariate: age). Prior to each ANCOVA test, a preliminary model ( interaction regression) was tested to determine if the slopes of the regression lines differed (homogeneity of slopes assumption; Sokal and Rohlf, 1981). The main significance test of the ANCOVA (homogeneity of y-intercepts) was performed when the parallelism of slopes assump- tion was met. Since differences in abundance affect the elevation of regression plots used to estimate mor- tality, spatial and temporal trends in mortality were evaluated by comparing slopes (y-intercept test not per- formed). Analysis of variance (ANOVA) was used to examine variability in density. Estimates of density were log -transformed to minimize heteroscedasticity. Results Environmental conditions Water temperature and salinity were recorded daily from August to December in 1994 and 1995. Tempo- ral variation in both parameters was pronounced and trends were similar between years (Fig. 2). Tempera- ture during the primary settlement period was vari- able (September-October) and ranged from 23.3° to 30.2°C in 1994 and from 22.7° to 31.6°C in 1995, re- spectively. Mean temperature during this period was similar between years 1995 (27.5°C) and 1994 (27.2°C). Salinity ranged from 25.3'?, to 37.07ff and from 25.3%r to 30.8%r in 1994 and 1995, respectively Similar to temperature, mean salinity was relatively similar between years: m.OVcc (1994) and 28.67« (1995). Peak values for both parameters occurred during the initial spawning period and declined for later spawning and settlement dates. Cohorts arriv- ing early in the season experienced high tempera- ture and salinity, whereas late-season cohorts were exposed to lower temperature and salinity (Fig. 2). No conspicuous differences in temperature or salin- ity were observed between sampling sites (ABl, AB2). Catch characteristics Overall, 1306 and 8144 red drum lai^vae and early juveniles (3-40 mm) were collected from the two sam- pling sites in the Aransas Estuary in 1994 and 1995. In both years, settlers were first detected in early September and peak densities of recently settled red drum (<10 mm) were present in early October (Fig 3). Larger postsettlers (>10 mm) were most abun- dant from mid-October to early November. Mean densities of recently settled red drum (<10 mm) ranged from 0.0 to 1.7/m' in 1994 and from 0.0 to 4.1/m- in 1995; densities of all postsettlement red drum (<40 mm) ranged from 0.0 to 3.0/m" in 1994 and from 0.0 to 4.2/m- in 1995 (Fig. 3). Maximum densities (per sled tow) observed in 1994 and 1995 in the Aransas Estuary were 3.4/m" and 11.5/m-, re- spectively. Interannual variation in postsettlement density during the two seasons (October-November) was small: 1.53/m2 (1994), 1.63/m'- (1995). No sig- nificant differences in density were detected between years or sites (ANOVA, P>0. 05). Although collection effort varied between years, length-frequency distributions of red drum from the Aransas Estuary were similar (Fig. 4 ). In both years, the smallest individuals collected were 3 mm and large recruits (>30 mm) were collected infrequently. Catch rates showed a steeply ascending left limb which peaked at approximately 8-9 mm, suggesting that recruitment to seagrass meadows was complete for these individuals. A long descending right limb characterized catches for individuals >10 mm. Age and growth Age-length relationships (age-length keys) were devel- oped for each year class and described by the following equations: Age = -17.05 -i- 42.25 LogSL (1994, «=249, r^=0.90, CV=8.9); Age = -11.43 -i-40.il LogSL (1995, n = UO, 7^=0.90, CV=7.5). Predicted ages of red drum (3-30 mm) ranged ft-om 6 to 45 d in 1994 and from 10 Rooker et al,: Variability in growth, mortality, and recruitment of Saaenops ocel/atus 585 34- 1995 32- ^^^-^ 1994 30- "--<::0\ g 28- \/-' V a; 26- 0 24- E ffl 22- 20- \ ^ — ^' 1995 294 277 26 0 24 3 22.1 20 8 V/ 18- 1994 28 5 27 1 26,2 25.2 24.3 23,2 Cohort A B C D E F 16" 1 1 1 1 1 1 1 1 1 1 1 1 1 1 1 1 1 1 1 1 1 1 1 1 1 1 r 1 ) 1 1 [ 1 1 1 1 ir 1 1 r 1 1 1 1 1 1 1 1 1 1 1 1 1 — ' 31 10 20 30 10 20 30 09 19 29 Aug Sep Oct Nov 38- 1995 36- -\ ,\,^\ 1994 34- \ d 32- \ CT3 28- ^ W--''^ A./ \ /\/ ' ''' /' '/ \ ^^'' \ V ' V '' ^ / f\r/ \, ,' \ 26- 1995 297 283 27 4 26 8 27,1 27 2 V / V ^^v"* 1994 30 9 28 9 28 5 28 0 27,4 26 4 [J Cohort A B C D E F 24 1 1 1 1 1 1 II 1 1 1 1 1 1 1 r 1 1 1 1 1 1 1 1 1 1 1 1 1 1 1 1 1 1 1 1 1 I 1 1 1 1 1 1 1 1 1 1 1 1 1 1 1 1 — 31 10 20 30 10 20 30 09 19 29 Aug Sep Oct Nov Date Figure 2 Water temperature and salinity records for the Aransas Estuary in 1994 and 1995. Mean temperature and salinity i30-d average) experienced by red drum cohorts lA-F) are given. to 48 d in 1995. Individuals from the most abundant size class (8-10 mm) were approximately 20-28 d. Age data were used to determine hatch dates and these data were used to partition individuals into 10-d cohorts. Instantaneous growth coefficients (g) ranged from 0.049 (4.8'7f/d) in 1994 to 0.051 (5.0%/d) in 1995 (Fig. 5) and differed significantly between years (ANCOVA.j- intercepts, P<0.001 ). Instantaneous growth coefficients were also estimated on a site-specific basis within the Aransas Estuary (Table 1). In 1994, growth rates at ABl were significantly higher than AB2 (ANCOVA, y-intercepts, P=0.015); however, no site effect was observed in 1995 (ANCOVA, y-intercepts, P=0. 440). The weight-specific growth coefficients (G) ranged from 0.159 to 0.162 (14.7-15.0%/d) and from 0.163 to 0.165 (15. 0-15. 2^^^) in 1994 and 1995, respec- tively (Table 1). Mortality Instantaneous mortality coefficients (Z) ranged from 0.134 (12.57c/d) in 1994 to 0.139 (13.0%/d) in 1995 and no significant interannual effect was observed (ANCOVA, slopes, P=0.737) (Fig. 6). Stage-specific mortality rates (cumulative mortality during the early postsettlement period: 12-d period, ca. 10-20 mm) ranged from 1.668 (81.17f ) to 1.608 (79.9'7c-) in 1994 and 1995, respectively. Spatial variation in mortality was also examined within each year by estimating mortality rates at the two sites within 586 Fishery Bulletin 97(3), 1999 the Aransas Estuary. Site-specific mortality rates in 1994 ranged from 0.129 to 0.141 and no significant difference was detected between sites (ANCOVA, slopes, P=0.741); however, in 1995 mortality rates were significantly lower in ABl (0.129) than in AB2 (0.193) (ANCOVA, slopes, P=<0.001). 45- ■ > ■ < 10 mm 10 mm 4.0- 3.5- 3.01 2.5- 2.0- 1.5- M 1.0- M 0.5- m 20 Sep 30 50] 4.5- 4.0- k 3.5- k 3.0- m 2.5- m 2.0- ■ 1.5- m 1,0- fl 0.5; 4 m 20 Sep 30 1994 30 29 1995 10 30 29 Date Figure 3 Mean density (sites pooled i per sampling trip of postset- tlement red drum collected from the Aransas Estuary in 1994 and 1995. Two size classes of postsettlement red drum (<10 mm SL, >10 mm SLi are shown. Table 1 Growth and mortality coefficients of postsettlement red drum collected from the Aransas Estuary in 1994 and 1995. The G:Z recruitment index is also shown. Year Site g G Z G:Z 1994 ABl 0.051 0.159 0.141 1.128 AB2 0.048 0.161 0.129 1.248 All sites 0.049 0.162 0.134 1.209 1995 ABl 0.052 0.165 0.129 1.279 AB2 0.049 0.163 0.193 0.845 All sites 0.051 0.165 0.139 1.187 Cohort-specific mortality rates were estimated for the 1995 year class (Fig. 7). Instantaneous mortal- ity coefficients for 10-d cohorts ranged from 0.106 (10.1%/d) to 0.265 (23.3%/d) and a significant tem- poral trend was detected (ANCOVA, slopes, P=0.004). Mortality rates were lowest for midseason cohorts (C, D) and highest for early and late cohorts (B, E), and individuals from the last cohort (E) experienced the greatest mortality. Stage-specific mortality rates (10-20 mm, 12 d) ranged from 1.272 to 1.524 (71.9- 78.2%) for midseason cohorts and 2.316 and 3.180 (90.1-95.8%) for early and late-season cohorts. G:Z Index Recruitment potential of different cohorts was evalu- ated by comparing G:Z ratios and temporal varia- tion was observed (Fig. 8). The ratio of G.Z was high- est for mid-season cohorts; 1.305 ( cohort C) and 1.565 (cohort D). In contrast, G:Z ratios of early and late- season cohorts (B, E) were <1, suggesting that these cohorts were losing biomass. Moreover, annual esti- mates of G.Z were calculated and ratios were >1 in both 1994 and 1995. Site-specific variation was also observed and, apart from site AB2 in 1995, G:Z ra- tios were always >1 (Table 1). Discussion Otolith-derived estimates of age indicate that red drum enter estuarine seagrass meadows after a short pelagic interval. Peak densities were observed for individuals 8-9 mm (20-24 d), suggesting that re- cruitment to seagrass meadows follows a planktonic period of approximately 20 d. In support of this find- ing, studies examining red drum during the plank- tonic phase have shown that the upper size range of red drum larvae in pelagic environments is 6-8 mm (Peters and McMichael, 1987; Comyns et al., 1989; Holt et al., 1989). In addition, Peters and McMichael (1987) sampled demersal habitats in Tampa Bay, Florida, and reported that red drum were first de- tected at 8 mm. Consequently, red drum appear to move from pelagic to demersal habitats (i.e. settle- ment) at approximately 8 mm. Once in seagrass meadows, length- and weight- specific growth of red drum increases rapidly with increasing size (exponential relationships). Length- specific growth rates of new settlers (20-40 d) in the Aransas Estuary averaged 0.58 and 0.62 mm/d in 1994 and 1995, respectively. Growth estimates in the Aransas Estuary are similar to rates reported by Peters and McMichael ( 1987 ) for red drum in Tampa Bay, Florida (0.58 mm/d). With size-at-age equations Rooker et al.: Variability in growth, mortality, and recruitment of Sciaenops ocellatus 587 200 100 ^ 1994 0 2 4 6 8 10 12 14 16 18 20 22 24 26 28 30 32 34 36 38 "5 1000 < uuu- 1995 800- 600- 400 200- 1 llli. n. .1 0 2 4 6 8 10 12 14 16 18 20 22 24 26 28 30 32 34 36 38 Standard length (mm) Figure 4 Length-frequency distributions of postsettlement red drum col- lected from the Aransas Estuary in 1994 (« = 1306) and 1995 (n=8144l. derived for red drum, we predicted that the length of a 40-d individual ranged from 20.2 mm ( 1994) to 18.2 mm ( 1995), which was similar to the predicted length of a 40-d individual collected in Tampa Bay (20.5 mm). Although seasonal trends in growth variation were not examined in this study, Rooker and Holt (1997) observed considerable variability among co- horts of newly settled red drum (0.5-0.8 mm/d), and growth was generally highest for midseason cohorts. Mortality rates of red drum were substantial and varied over spatial and temporal scales. Interannual estimates of mortality were similar and losses ranged from 12.5 to 13.0%/d (Z=0.13-0.14). Consequently, cumulative mortality during the settlement and early postsettlement period (12-d period) was relatively high; only about 20'7f of the initial settlers survived to 40 d. Although no estimates of mortality exist for red drum larvae and early juveniles, these mortality rates are comparable to values reported for other marine fishes during the early life stage. Houde (1987) reported instantaneous daily mortality coef- ficients for five species of marine fishes during the larval stage (first feeding to late-stage larvae) and 30- SL = 2 957e''«9la9e) r2 = 0.90 / O / 20- OCpyo O C^^" 10- ^\^^° ^^^^ 1994 s: 10 20 30 40 50 " Age (d) 1 Standard l( SL = 2 368e<'<'5i (ags) r^ = 0,90 / 20- 0 / oo o 0 /o o q/o CX) / _^^o 10- ooet> oo o oog--^^ ^ — -"o o 1995 10 20 30 40 50 Age (d) Figure 5 Size-at-age relationships of postsettlement red drum collected from the Aransas Estu- ary in 1994 (/!=249) and 1995 (n = 140). Ex- ponential equations are given. predicted values ranging from 0.04 to 0.18 (4-16%/d). In addition, Houde and Zastrow ( 1993) summarized natural mortality data for teleost larvae from a vari- ety of sources and reported that the mean instanta- neous daily mortality rate of larvae in estuarine habi- tats was 0.266 (23%/d). Also, McGurk (1986) devel- oped a general mortality equation for marine fish lai-vae (based on dry weight), and using his equation we predicted the mortality rate of individual red drum 8-10 mm (ca. 750-1500 |ig dry wt) would be approximately 0.05-0.11 (4.9-10.4%/d). Although interannual patterns of mortality were insignificant, variability within a year was relatively high and a seasonal trend was evident. Cohort-spe- cific estimates indicated that mortality rates were lowest for red drum produced in the middle of the spawning season ( 107c/d), when densities within the seagrass meadows reached maximum levels. Con- versely, mortality rates for early and late-season co- 588 Fishery Bulletin 97(3), 1999 LoQe N, = 10.786 • 0 139(age) LoQeN, = 8.350-0 134(age) r2 = 0.84 ^ 1995 A 1994 20 30 40 50 Age (d) Figure 6 Regression plots of Log^, abundance on age of postsettlement red drum collected from the Aransas Estuary in 1994 and 1995. Re- gression equations are given. 7] 6 5- 4- 3- 2 H 0 20 7- 6 5 4 3 2-1 1 0 Logg N, = 11 609-0 193(age) r2 = 0.87 cohort B 30 40 50 Lege N, = 6.169 -0.106(age) r2 = 0.34 cohort D 7- 6 5 4 3- 2- 1- 0 20 7 6- 5 4 3 2 1 0 LoggN, = 8.503-0 127(age) r2 = 0.78 cohort C 30 cohort E 20 30 40 50 20 Age (d) 30 Figure 7 Regression plots of Log, abundance on age for 10-d cohorts of postsettlement red drum collected from one site (AB2) in the Aransas Estuary in 1995. Regression equations and plots are shown for cohorts B, C, D, and E. horts were approximately twofold higher, approach- ing 259(^/d for the late-season cohort. Seasonal fluc- tuations of this magnitude have been reported in several studies on marine fish larvae and attributed to changes in environmental conditions (e.g. tempera- ture, prey availability) which can directly or indi- rectly influence mortality (Rutherford and Houde, 1995; Secor and Houde, 1995). Temperature is often implicated as a critical fac- tor in recruitment because it has the potential of in- fluencing growth rates and causing episodic mortal- ity during early life (Houde and Zastrow, 1993; Houde, 1996). Temperature has been shown to be a primary determinant of growth variation in labora- tory and wild populations of red drum (Holt et al., 1981; Lee et al., 1984; Rooker et al., 1997). More- over, Rocker and Holt (1997) determined that sea- sonal trends in growth variation for red drum were related to temperature. Mean temperatures experi- enced by successive cohorts declined throughout the season; however, growth rates were highest for mid- season cohorts (20 Septem- ber-10 October) and lowest for early and late cohorts. Conversely, mortality rates were lowest for mid-season cohorts and highest for early and late cohorts. Growth and mortality data appear to be best described by cur- vilinear relationships (qua- dratic function), suggesting that there may be an opti- mal temperature range at which growth and survival of red drum are enhanced and that the midpoint of this range occurs at approxi- mately 26' 'C. Variability in other biotic and abiotic factors may also be responsible for observed trends in mortality. Similar to other marine fishes, growth and survival rates of red drum larvae are associ- ated with prey availability (G.J. Holt, unpubl. data). Since spatial and temporal variation in prey density (i.e. meiofauna) are common in estuarine habitats in south Texas (Montagna and Kalke, 1992), seasonal trends in prey abundance may 40 50 LoggN, = 11 918-0.265(age) r^ = 0.91 40 50 Rooker et al.: Variability In growth, mortality, and recruitment of Saaenops ocellatus 589 C D E Cohort (1995) 1994 1995 Year Figure 8 Recruitment potential (G;2 ratios) of 10-d cohorts of postsettlement red drum collected from the Aransas Estuary in 1995. Interannual estimates of G.Z are given. have affected early life survival. Conversely, preda- tor fields also vary spatially and temporally and sur- vival may be a function of predator abundance. Postsettlement red drum are exposed to a suite of predators in the seagrass meadows and the abun- dance and distribution of these predators are highly variable (senior author's unpubl. data). Rooker et al. (1998a) examined predation rates on newly settled red drum in experimental mesocosms and demon- strated that predators inhabiting these seagrass meadows, are capable of consuming large numbers of red drum (Z=2-7'^/[h • predator] ). Consequently, fluctuations in prey and predator densities appear critical to the survival of red drum and the addition of this information to future studies should enhance our understanding of early life mortality. Growth and mortality estimates were combined to evaluate the relative recruitment potential iG.Z) of red drum cohorts. According to Houde and Zastrow's (1993) review of 188 species, G:Z ratios for marine fish larvae are generally less than 1.0 (mean G.Z=0.89) and larvae inhabiting estuarine ecosys- tems tend to have low G;Z ratios (range: 0.34-0.82). However, recent work in Chesapeake Bay on striped bass {Morone saxatilis) has demonstrated that G:Z ratios can be >1 in estuarine ecosystems (Ruther- ford and Houde. 1995; Secor and Houde, 1995). Simi- larly, G:Z ratios for postsettlement red drum in the Aransas Estuary were >1 in both 1994 and 1995 (1.21 and 1.19, respectively) and, as a result, gross growth efficiency was positive and the recruitment poten- tial of 1994 and 1995 year classes appears to have been favorable. A high degree of correspondence in G:Z ratios be- tween years was present, suggesting that conditions for growth and survival were similar in 1994 and 1995. Although the assessment of physical and bio- logical conditions was limited (no data on prey avail- ability or predator fields), parameters measured in this study (e.g. temperature, salinity, settlement size, and density ) were relatively equal between years. In contrast, G:Z ratios of 10-d cohorts and the environ- mental conditions experienced by these individuals were highly variable within a single season. The G:Z ratios were highest (1.3-1.6) for midseason cohorts and settlement densities at these times were at maxi- mum levels. Thus, it appears that physical and bio- logical conditions in the seagrass meadows were op- timal for growth or survival, or for both. The oppo- site trend was observed for individuals arriving early and late in the season. Because G:Z ratios of indi- viduals from these cohorts were low (0.6-0.9), we postulated that nursery conditions experienced by early and late-season cohorts do not favor early life survival (low recruitment potential). Acknowledgments We thank Kathy Binney, Sharon Herzka, Patti Pickering, Cameron Pratt, and Andy Soto for pro- viding assistance in the laboratory and field. The manuscript benefited from the comments of three anonymous reviewers. This work was supported by grants from the Texas Higher Education Coordinat- ing Board Advanced Research Program (grant 003658-392 ), Texas A&M Sea Grant Program (grant NBR;RIF65), Sid W. Richardson Foundation, and fellowships to J.R.R (E.J. Lund, Julian C. Barton). Literature cited Bailey, K. M., and E. D. Houde 1989. Predation on egg and larvae of marine fishes and the recruitment problem. Adv. Mar Biol. 25:1-83. Baltz, D. M., C. Rakocinski, and J. W. Fleeger. 1993. Microhabitat use by marsh-edge fishes in a Louisi- ana estuary. Environ. Biol. Fish. 36:109-126. Campana, S. E., and J. D Neilson 1985. Microstructure of fish otoliths. Can. J. Fish. Aquat. Sci. 42:1014-1032. Comyns, B. H., J. Lyczkowski-Shultz, D. L. Nieland, and C. A. Wilson. 1991. Reproduction of red drum, Sciaenops ocellatus, in the northcentral Gulf of Mexico: seasonality and spawner biomass. U.S. Dep. Commer, NOAA Tech Rep NMFS 95:17-26. Comyns, B. H., J. Lyczkowski-Shultz, C. F. Rakocinski, and J. R. Steen Jr. 1989. Age and growth of red drum larvae in the north-cen- tral Gulf of Mexico. Trans. Am. Fish. Soc. 1 18: 159-167. Cushing, D. H. 1975. Marine ecology and fisheries. Cambridge Univ. Press, Cambridge, England, 278 p. 590 Fishery Bulletin 97(3), 1999 Frederick, J. L. 1997. Post-settlement movement of coral reef fishes and bias in survival estimates. Mar. Ecol. Prog. Ser 150:65- 74. Fuiman, L. A., and A. E. Magurran. 1994. Development of predator defences in fishes. Rev. Fish. Biol. Fish. 4:145-183. Holt, G. J., R. Godbout, and C. R. Arnold. 1981. Effects of temperature and salinity on egg hatching and larval survival of red drum. Sciaenops ocellata. Fish. Bull. 79:569-573. /Holt, S. A., G. J. Holt, and C. R. Arnold. 1989. Tidal stream transport of larval fishes into non-strati- fied estuaries. Rapp. P.V. Reun. Cons. Int. Explor Mer 191:100-104. Houde, E. D. 1987. Fish early life dynamics and recruitment variability. Am. Fish. Soc. Symp. 2: 17-29. 1996. Evaluating stage-specific survival during the early life of fish. In Y. Watanabe. Y. Yamashita. and Y. Oozeki (eds.). Survival strategies in early life stages of marine resources. AA Balkema, Rotterdam, p 51-66. Houde, E. D., and C. E. Zastrow. 1993. Ecosystem- and taxon-specific dynamic and energet- ics properties of larval fish assemblages. Bull. Mar. Sci. 53:290-335. Jones, C. 1986. Determining the age of larval fish with the otolith increment technique. Fish. Bull. 84:91-103. Lee, W. v., G. J. Holt, and C. R. Arnold. 1984. Growth of red drum larvae in the laboratory. Trans. Am. Fish. Soc. 113:243-246. McGurk, M. D. 1986. Natural mortality of marine pelagic fish eggs and larvae: role of spatial patchiness. Mar Ecol. Prog. Ser 34:227-242. Montagna, P. A., and R. D. Kalke. 1992. The effect of freshwater inflow on meiofaunal and macrofaunal populations in the Guadalupe and Nueces Estuaries, Texas. Estuaries 15:307-326. NOAA (National Oceanic and Atmospheric Administration). 1991. Our living oceans: First annual report on the status of U.S. living marine resources. U.S. Dep. Commer., NOAA Tech. Memo, NMFS-F/SPO-l:47-48. Peters, K. M., and R. H. McMichael Jr. 1987. Early life history of the red drum Sciaenops ocellatus (Pisces: Sciaenidae) in Tampa Bay, Florida. Estuaries 10:92-107. Ricker, W. E. 1975. Computation and interpretation of biological statis- tics offish populations. Bull. Fish. Res. Board Can. 191, 382 p. Rooker, J. R. 1997. Early life history of red drum iSciaenops ocellatus) in subtropical seagrass meadows: patterns of condition, growth, and mortality. Ph.D. diss., Univ. Texas at Aus- tin, Austin, TX, 192 p. Rooker, J. R., and S. A. Holt. 1997. Utilization of subtropical seagrass meadows by newly settled red drum {Sciaenops ocellatus): patterns of distri- bution and growth. Mar Ecol. Prog. Ser 158:139-149. Rooker, J. R., G. J. Holt, and S. A. Holt. 1997. Condition of larval and juvenile red drum ISciaenops ocellatus) from estuarine nursery habitats. Mar. Biol. 127:387-394. Rooker, J. R., G. J. Holt, and S. A. Holt. 1998a. Vulnerability of newly settled red drum (Sciaenops ocellatus) to predatory fish: is early survival enhanced by seagrass meadows? Mar Biol. 131:14.5-151. Rooker, J. R., S. A. Holt, M. A. Soto, and G. J. Holt. 1998b. Post-settlement patterns of habitat use by sciaenid fishes in subtropical seagrass meadows. Estuaries 21:31.5-324. Rutherford, E. S., and E. D. Houde. 1995. The influence of temperature on cohort-specific growth, survival, and recruitment of striped bass, Morone saxatilis, larvae in Chesapeake Bay. Fish. Bull. 93:315- 332. Secor, D. H., and E. D. Houde. 1995. Temperature effects on the timing of striped bass egg production, larval viability, and recruitment potential in the Patuxent River (Chesapeake Bay). Estuaries 18:527- 544. Sokal, R. R., and F. J. Rohlf. 1981. Biometry, 2nd edition. W. H. Freeman, San Fran- cisco, CA, 859 p. Swingle, W. E. 1990. Status of the commercial and recreational fishery. In Red drum aquaculture, p. 22-24. Texas A&M Sea Grant Program report TAMU-SG-90-603, College Station, Texas. Werner, E. E., and J. F. Gilliam. 1984. The ontogenetic niche and species interactions in size- structured populations. Ann. Rev. Ecol. Syst. 15:393-425. 591 Abstract.— Relative abundance of sablefish, Anoplopoma fimbria, in the waters off Alaska has been measured annually since 1979 with longline sur- veys. These extensive surveys provide precise measures of relative abun- dance. An age-structured model was fitted to the longline survey data to es- timate absolute abundance. Estimates of recent exploitation rates for fully se- lected ages averaged IC/r. Monte Carlo simulations of the age-structured model indicated that absolute abundance of Alaskan sablefish can be estimated re- liably if age selectivity is asymptotic and not dome-shaped. Abundance esti- mates were reliable even when only length data and no age data were avail- able. Dome-shaped selectivity gave bi- ased and less precise estimates, prob- ably owing to a parameter interaction between catchability and the shape of the selectivity function. Estimation of sablefish, Anoplopoma fimbria, abundance off Alaska with an age-structured population model Michael F. Sigler Auke Bay Laboralory, Alaska Fisheries Science Center 11305 Glacier Hwy, Juneau, Alaska 99801-8626 E-mail address Mike Sigler@noa3.gov Manuscript accepted 18 August 1998. Fish. Bull. 97:591-603 (19991. Sablefish, Anoplopoma fimbria, is a long-lived species that inhabits the northeast Pacific Ocean and Bering Sea. This species supports a fishery in Alaskan waters, with catches ranging from about 10,000 to 35,000 metric tons (t) during the last two decades (Fig.l). The fish- ery mostly uses longline gear and primarily occurs on the upper con- tinental slope, which is inhabited by adult sablefish (Fig. 2). Previously the fishery became compressed year-round during the mid-1980s to 10 days in some areas, until 1995, when management switched to indi- vidual fishing quotas and an eight- month season. The delay of the start of the season from 1 January to 1 April and later 15 May accompanied the compressed fishery season. In Alaska, relative abundance of sablefish has been best measured by annual longline surveys since 1979. Longline surveys are preferred over trawl surveys for assessing sable- fish because the longline surveys generally cover the areas that adult sablefish inhabit, namely the upper continental slope. The longline sur- veys have occurred between early May and late September. The sur- vey and fishery generally take place in the same area, the upper conti- nental slope but the fishery gener- ally takes place over a narrower depth range. Trawl surveys have also been conducted in Alaska, but compared with the longline surveys, they have not sampled as deeply and have covered fewer years. In stock assessments for Alaskan sablefish prior to 1996 (Fujioka, 1995; Lowe, 1995), the limited trawl data were used to convert relative abundance from the longline survey to absolute abundance by calibra- tion to the trawl (Rose, 1986), and the population was modeled by us- ing a delay-difference analysis (Kimura, 1985). The formulation of the delay-difference analysis ap- plied to Alaskan sablefish was modi- fied to provide annual recruitment estimates, with the assumption that trawl surveys measure absolute abundance (Fujioka, 1995; Lowe, 1995). The Alaskan sablefish stock assessment was like other Alaskan groundfish stock assessments, which historically have assumed that trawl surveys measure abso- lute abundance (Alverson and Pereyra, 1969). This assumption probably is wrong because fish can escape under the net (Engas and Godo, 1989a) and be herded by the bridles (Engas and Godo, 1989b). The motivation for the current study was to assess Alaskan sable- fish without this assumption, rely- ing only on longline survey esti- mates of relative abundance. In the current study, I estimated absolute abundance for Alaskan sablefish with an age-structured population model, evaluated the estimation approach with Monte Carlo simulations, and investigated interactions of model parameters. Delay-difference analyses use an- nual abundance estimates and catches, whereas age-structured analyses additionally use annual 592 Fishery Bulletin 97(3), 1999 age or length data. The additional information al- lows explicit tracking of cohorts, but many more pa- rameters must be estimated. Fitting population models to survey and fishery data has long been used to estimate absolute abun- dance (Pope and Shepherd, 1985; Deriso et al., 1985) and is one method of estimating absolute abundance along with direct surveys, mark-recapture experi- ments and depletion experiments. Abundance esti- 'mation with a population model works by tracking population additions and losses and inferring abso- lute abundance from the manner in which catch af- 40 o 10 1979 1981 1983 1985 1987 Year 1989 1991 1993 1995 Figure 1 Reported fishery catches (thousands metric tons) from 1979 to 1995. w \J. *^ "^ Russia \ "( ^_y^ Alaska Canada -60N \ -50N 200 m contour of continental slope 160W 150W 1 ,1. . 140W 1 170E 180 1 1 1 70W i... fects the survey index and age and length composi- tions. In simple terms and aside from other losses, if harvesting 500 individuals decreases a cohort's in- dex 10%, then the cohort's initial abundance was 5000 individuals. Estimation depends on having an abundance index precise enough to detect cohort re- ductions by the fishery. The population model presented here takes into account the following unusual attributes of the avail- able data. Length compositions are available from the longline surveys for all years analyzed, compared with infrequently available age compositions (Table 1). Not many Alaska-wide sablefish fishery data are available, consisting only of total catches and five years of fishery lengths, but no ages (Table 1) . Methods Model structure and estimation method The analysis generally follows the approach de- scribed by Kimura ( 1990 ) for a separable age-struc- tured population model. Let y, a , and / be the year, age, and length indices, respectively. c , = observed fishery catch in numbers; /, = observed survey abundance index in numbers; p = obsei"ved survey proportion at age; p ,1 = observed survey proportion at length; jj = Ij s = the exploitation fraction of age-a fish during yeary which is separable into s , the selectivity for age-a fish, and jj ,, the exploitation fraction for fully vulnerable ages; A^ A^. .A^. X<=^v the total number at age; the exploitable (fishable) number at age; and the exploitable number. Figure 2 Map of southern Alaskan waters. Adult sablefish inhabit the upper continental slope seaward of the 200-m depth contour. Starting with initial cohort size, N^,^^, natural deaths and observed catch were removed to compute next year's cohort size, A^ , , = ( 1-t/ ) N e-^. A discrete fish- v+l.a + l ^ \(i ya ery was modeled, c'y = ilyN\, because the fishery was short for recent years except 1995. The modeled discrete fishery gener- ally matches the timing of the observed dis- crete fishery and occurs near the mid-point of the earlier year-round fishery Sigler: Estimation of abundance of Anop/opoma fimbria off Alaska 593 Table 1 Data sets used in the age-structured model of Alaskan sablefish. Length data are specified by sex. Data component Data aggregation Years of data Longline survey relative abundance Longline survey male lengths (cm FL) Longline survey female lengths (cm FL) Longline survey age Fisherv total catch 40-41. 42-43 62-63, 64-69. 70-74, ..., 95-99 40-41, 42-43 68-69, 70-74, 75-79, ..., 95-99 2, 3, 4, 5, 6, 7, 8, 9-10, 11-15, 16+ years 1979-95 1979-95 1979-95 1983, 87, 89, 91, 93 1979-95 The modeled discrete fishery misrepresents the duration of the earher year-round fishery, but tests with simulated data comparing a continuous fishery showed little difference in estimated abundance. The assumption of a discrete fishery simplified catch ac- counting and reduced model convergence time, which was important for the Monte Carlo simulations. This formulation limits the maximum exploitation rate for an age class to equal the selectivity for that age. This maximal rate would be achieved if the maxi- mally selected age were completely harvested, with the exploitation rate equal to 1.0. This potential prob- lem is not a real one because exploitation rates never approach such high levels for Alaskan sablefish. Selectivity was described by the "exponential-lo- gistic" function (Thompson, 1994), U-yJ i-y PriA,,„-a> \ [1 + e litA,,"- The "exponential-logistic" function is flexible, allow- ing both asymptotic selectivity when selectivity in- creases with age to an asymptote, and dome-shaped selectivity when selectivity increases with age to a maximum, then decreases for older fish. The expo- nential-logistic function automatically scales maxi- mum selectivity to 1.0 and reduces to asymptotic selectivity as the parameter gamma (7) approaches zero. When y= 0, the parameter A^g is the age where 500r of the population is vulnerable and /J is the slope of the function at Ag^. When 7 > 0, then A^g and P lose their biological meaning, because A^q no longer represents the age at 50'^^ vulnerability. Selectivity is assumed equal for the fishery and survey. Both the fishery and survey mostly use longline gear and cover the same area. Their similar length frequencies support this assumption (Fig. 3). The fisheiy length data were not incorporated into the model because of this similarity and because few years of fishery length data were available. Age data were aggregated over adjacent ages (Table Das suggested by Derisoetal.( 1989) because sablefish are difficult to age, especially those older than eight years (Table 6 in Kimura and Lyons, 1991). Ages greater than eight years were not pooled into a single class because females continue to grow, though slowly, after eight years. This growth information was needed in the model to convert ages to lengths. Length data also were aggregated (Table 1 ). Another way to deal with ageing error is to include a matrix of misageing probabilities in the age-structured model (Kimura, 1990), but this approach was not used be- cause the misageing probabilities were unknown. Estimated data values were computed from the parameter estimates. The estimated abundance in- dex is I^ = qN[. ,where q is survey catchability. Quan- tities estimated from this model and used in the model fitting algorithm are denoted with "hats." The estimated age compositions are N Pva / , V An age-length transition matrix, L = (/^^p, also was calculated from the sui-vey age data, where /^^, is the probability that a sampled fish of age a will be of length /. The age-length transition matrix is assumed constant over time. Probabilities were computed by region and year, then pooled as an average weighted by sample size. Estimated age compositions were converted to estimated length compositions with the age-length transition matrix. Pv/ ^P.Jal The parameters were estimated by maximum like- lihood by assuming multinomial errors for age and length data and log-normal errors for catch data (Fournier and Archibald, 1982) and by using the quasi-Newton algorithm implemented in Microsoft Excel. Assuming that the effective sample sizes are the same for age and length samples and that the 594 Fishery Bulletin 97(3), 1999 0 0 ll I f I I I I M I M ' I I M ri-i 40 46 52 58 64 70 85 40 46 52 58 64 70 85 0,0 40 46 52 58 64 70 85 40 46 52 58 64 70 85 0.3 -I 40 46 52 58 64 70 85 0.3 1 0.2 0.1 0.0 -I M ^ t 40 46 52 58 64 70 85 Length class (cm) Figure 3 Survey ( ) and fishery t ) length compositions for females. Male lengths also are available, but only female lengths are displayed for brevity. Alaska-wide fishery length data were available only since 1990. age and length samples are independent, then the log-likelihood is Pxl ^ = X Py log ^^ + X Py' ^°^ ya Pya yi Pyl -^'Y\].og{Iy)-\Qg(qN^)\ + constant, where A is the ratio of the effective multinomial sample sizes, either age or length, to the variance of the log-transformed abundance index (Deriso et al., 1985). Thus A is a weighting factor that adjusts the relative influence of the abundance index and age and length components of the likelihood. The range of A from 0.02 to 10 was tested. The value A = 1 ap- peared the most reasonable and is used henceforth. This value was chosen on the basis of fit of each data type and the effect on the estimated recruitment values as A was varied. As A increased from 0.02 to 10, the fit of the index data improved only at the expense of the fit of the length data, but the relative sizes of the annual recruitment estimates were un- affected for A < 2. Increasing A substantially improved the fit of the index data until A was from 1 to 2, when improvement slowed. Thus a value of A between 1 and 2 seems reasonable, with A = 1 chosen because the fit of the age data was best at that point. In prac- Sigler: Estimation of abundance of Anoplopoma fimbria off Alaska 595 tice, A = 1 implies an effective sample size for the age and length samples of 100 if the CV of the sur- vey index is 10%. This CV is reasonable according to a statistical analysis of the survey index, which found that a 10% interannual change in the survey index typically was statistically significant at the P= 0.05 level (Sigler and Fujioka, 1988). The reasonableness of the effective sample size is harder to determine because the age and length samples are composites of multinomial samples from survey stations with underlying geographic variation in age and length compositions, so that the effective sample size is less than the true sample size. In the above expression log (p ,^j) usually found in the log-likelihood for the multinomial distribution was replaced with log ( p ^J p .J- This replacement removes a "nonsignificant" portion of the likelihood and makes it easier to ex- amine the model fit and probably somewhat improves numerical performance (Kimura, 1990). Assuming M and A are known, this model contains Y + A + 2 parameters: recruitment, TV^j ..., A^ ., the initial age composition, N q^^, ..., N ^^, and the selec- tivity parameters, A^q, /3, and y are unknown. The quantities and /ij, ..., /}y are functions of the ob- served survey indices and observed catches and of the parameter estimates. Setting clL/dq equal to zero and solving for q gives q = exp 1'°^^ N. Y The /j J , . . . , jUy are computed from /; ~ Ni ' by treating reported catches as exact. Although clearly there will always be some error in the reported catch, I concluded that this approximation generally was reasonable given the comprehensive system for tracking the Alaskan sablefish catch, which includes processor reports, fish receipts, individual fishing quota landing reports, and observer coverage. I esti- mated log-parameters rather than parameters on the original scale to improve reliability in the estima- tion process (Kimura, 1989, 1990). An allowable biological catch (A5Cgg) was calcu- lated from a 1-year future projection. The constant fishing mortality, F^^r, , was applied, which reduces the exploitable population to 40% of the unexploited state (Clark, 1991). A 1 -year projection of recruitment was forecast as the average of recent estimated re- cruitments. The two most recent recruitments were not used in the average; their estimates, based on only one or two years of data, are unreliable. Validation of estimation method I used Monte Carlo simulation for model validation (Kimura, 1989; Press et al., 1989) to verify that the age-structured analysis provides reasonable results. The steps (Press et al., 1989) are as follows: 1) Fit the model using real data; 2) take the estimated pa- rameter values as the true values for the simulation; 3 ) calculate the expected data based on these param- eter values, then simulate a new data set by adding a particular error to the expected data based on an observed mean square error ( MSE ) or any hypotheti- cal value; and 4) estimate the model parameters for the simulated data. If the resulting parameter esti- mates and the true parameter values are similar, then the estimation method is to some extent vali- dated. This approach assumes that the model fits the simulated data perfectly except for random er- ror. If the true errors were of larger magnitude or arose from a different source than that assumed in fitting the model, for example in violating the as- sumptions of equal fishery and survey selectivity and constant growth rates, then estimation performance with simulated data overstates the true model per- formance. Log-normal error (CV=0.10 on the original scale) was added to the expected abundance index, and multinomial error (n =200 1 was added to the expected age and length data. These values were based on variability of the model residuals (difference between the observed and expected data). The log-transformed abundance indices are assumed to be independent, normally distributed random variables, and a "mi- nor dilemma" (Kimura, 1989) arises on whether to simulate the abundance indices so that they are un- biased on the original or the logarithmic scale. I chose to simulate the abundance indices so that the sur- vey index was unbiased on the original scale. Age data are often limited in availability, whereas length data are nearly always available. Age data are more desirable than length data for estimating age structure because the correspondence between age and length is not one-to-one. Age data are com- monly inaccurate: some individuals, especially older fish, are misassigned to ages around the true age. Thus, inaccurate age data and length data are simi- lar in that true age does not uniquely correspond to assigned age or length. Parameters were estimated from simulated data for both accurate and inaccu- rate age data, the latter also being equivalent to length data. For accurate age data, 100% offish of simulated samples were correctly assigned and cat- 596 Fishery Bulletin 97(3), 1999 egorized as |2, 3, ..., 15, 16+) . For inaccurate age or length data, 50% were correctly assigned to the true age, 'ZOVc to -1 year, 20% to +1 year, 5% to -2 years, and 5% to +2 years, then categorized as 12, 3, ..., 8, 9+). Survey selectivity may affect parameter estima- tion. Bence et al. ( 1993 ) found that biomass estimates were more accurate and precise for a population model with data from a survey with asymptotic se- lectivity than from one with dome-shaped selectiv- ity. Therefore, both asymptotic and dome-shaped se- lectivity were compared. The dome-shape was cho- sen such that the availability of the last age group was 0.5. A measure of the variability of biomass estimated in the simulations is the MSE computed from pa- rameter estimates and their "true" values. Mean square error (MSE I was converted to coefficient of error (CE), defined as the "true" value divided into the square root of MSE (Kimura, 1990). Twenty-five to fifty replicate simulations were completed for each scenario. Results Model results for the original data The model fitted all the original data well: abundance index ( Fig. 4 ), age data ( Fig. 5 ), and length data ( Fig. 6 ). The estimate of survey catchability appears good; the likelihood profiled over a range of catchabilities shows a distinct, regular curvature (Fig. 7). Esti- mated exploitable biomass for 1995 was 181,000 t (Fig. 8), and projected ABCgg was 19,600 t. Estimated biomass decreased from a peak in the mid-1980s. The 1400 1200 ■/\- rvey index (numbers) a> CD Q 8 8 8 ^^ >\ y \.^ 400 200 1979 1981 1983 1985 1987 1989 1991 1993 1995 Figure 4 Observed ( ) and expected i - - -- ) survey index (num- bers) from 1979 to 199.5. peak is attributed to strong recruitment in the late 1970s; recruitment has decreased in recent years (Fig. 9). Estimates of recent exploitation rates for fully selected ages average 107f (Fig. 8), which is near the exploitation rate equivalent to F^^y., the current reference point for sablefish management in Alaska. The shape of the estimated selectivity curve was as- ymptotic (i.e. Y=0). The estimate of natural mortality for sablefish is uncertain; therefore its effect on abundance estima- tion was examined. An important part of this exami- nation was to analyze the interaction between M and other key parameters. Model parameters were esti- mated for several fixed values of natural mortality around M = 0.10 (Table 2). The log-likelihood is not maximized at M = 0. 10, and there is a slightly higher value for M = 0.12 (panel 1). Catchability was smaller and exploitable biomass was larger for larger M ( panel 2); biomass was larger to account for more natural deaths. The fishable fraction of the total population .N. /I^vo V a was smaller for larger M (panel 3) because the fish recruited later (panel 4). Asymptotic selectivity was estimated whatever the value of M (panel 5). A5Cgg was larger for larger M( panel 6) because exploitable biomass was larger and the fishing rate, F^^^., in- creases with M. Natural mortality and survey catchability can af- fect abundance estimates; therefore their effect was examined. Model parameters were estimated for sev- eral fixed values of g and M (Table 2). For each fixed M, the approach was to fix q at values near the esti- mated q. The results are likelihood profiles of q for each fixed M (panel 1). Given M, exploitable biomass was smaller for larger q (panel 2). The effect of g on the fishable fraction was more complicated. Given M, the fishable fraction was smaller for the fixed q that was less than the estimated q (panel 3) because older fish were less vulnerable to the fishery (panel 5). Given M, the fishable fraction was less for the fixed q that was greater than the estimated q be- cause fish recruited later (panel 4). Selectivity was asymptotic and could not increase above 1.0, forcing the decreased fishable fraction to occur by means of a later recruitment age. ABCgg was larger for smaller values of g (panel 6) because exploitable biomass was larger (panel 2). Model results for the simulated data Biomass estimates were unbiased for simulations based on a survey with asymptotic selectivity, when Sigler: Estimation of abundance of Anop/opoma fimbria off Alaska 597 0.3 1 23456789 11 16 3456789 11 16 2 3 4 5 6 7 9 11 16 23456789 11 16 23456789 11 16 Age class Figure 5 Observed ( i and expected ( i age compositions. Data were available only from the 1983. 1987. 1989. 1991. and 1993 surveys. See Table 1 for age-class definitions. age and length data were used (Fig. lOA). Estimated error (CE [exploitable bioniass]) was about 0.08, in- creasing for the later years to about 0.12 (Fig. Ill, values similar to the assumed CV for the abundance index of 0.10. These values all imply that sablefish abundance could be estimated given the type and quality of data assumed in these simulations. These simulations match my evaluation of the actual data. Biomass estimates also were unbiased for asymptotic selectivity, no matter whether age or "length" data were used (Fig. 10, B and C); CE (exploitable bio- mass) ranged from 0.10 to 0.15 (Fig. 11). In contrast, biomass estimates were positively biased by 149^ to 179^ for dome-shaped selectivity (Fig. lOD) and CE ( exploitable biomass ) ranged from 0.40 to 0.50 ( Fig. 11). Thus, estimating abundance from a survey with as- ymptotic selectivity appears reasonable; biomass estimates derived from a survey with dome-shaped selectivity can be estimated, but may be biased and less precise. Discussion Comparison of age-structured and delay-difference analyses The estimates of sablefish exploitable biomass from age-structured analysis and delay-difference analy- sis are similar (Fig. 12). Both analyses show exploit- able biomass rising from about 170,000 t in 1979, peaking at about 420,000 t around 1986, and falling since then to about 200,000 t in 1995. The trend from the sablefish delay-difference analysis is less smooth 598 Fishery Bulletin 97(3), 1999 Males 40 46 52 58 64 80 95 0,3 . 1985 1 ■1 0.2 A 0.1 / 'l n n .--^' 1 > 1 1 I 1 40 46 52 58 64 80 95 40 46 52 58 64 80 95 Females 0.0 i-^ 40 46 52 58 64 70 85 40 46 52 58 64 70 85 Length class (cm) Figure 6 Obsen-ed i ) and expected ( i length compositions. Data were collected during all annual surveys from 1979 to 1995. but only a subset is displayed for brevity. See Table 1 for length-class definitions. because the estimates are linear functions of the sur- vey abundance index. Exploitable biomass follows every up and down in the abundance index. All mea- surement error is reflected in the recruitment esti- mate. The trend from the age-structured analysis is a complicated function of the abundance index and the age and length data and smooths annual changes in the abundance index. The simulations in this paper imply that the age- structured model provides reliable estimates of sable- fish absolute biomass. The assumption that trawl surveys provide absolute biomass estimates required for the sablefish delay-difference analysis can be dropped. Starting with the 1997 fishing year, the age- structured analysis has supplanted the delay-differ- ence analysis in the stock assessment for estimating biomass(Fujioka et al., 1996). Comparison to other estimates of sablefish abundance Sablefish absolute abundance was estimated in two other studies, but the estimates were unreasonable. Sablefish longline catchability was estimated by depletion experiments (Clausen et al., 1997). The computed longline catchabilities were applied to the sablefish longline survey results for the Gulf of Alaska; the resultant biomass estimates of 50,000- Sigler: Estimation of abundance of Anoplopoma fimbria off Alaska 599 6 7 Catchability coefficient 8 9 10 11 12 -1 78 -1 79 T3 •1 80 ■ CD O -1 81 - -182 / -1 83 / •1.84 - Figure 7 \ Log likelihood versus catchability coefficient. 0 12 A 400 /\/ ^ / ^^ 0 10 E ** / ^^^ ^N^ ■o S 300 i' / ■■ 0 08 \ ' / , £ \ " / * ^ o s V / \ 0,06 % i 200 - \ / \ D o--9 1.000 500 X= 1-16 1 0.75 0.5 0.25 0 0 0.2 0,4 0.6 0.8 1 1,500-1 1 ,000 500 B Balistes vetula ,...o-9 tc>o--o-;a"-o--D--o--a--o-°- ;v= 1 11 1 -0.25 0 Yield without reserve ORP Yield with ORP 0 0.2 0.4 0.6 0.1 1 1 2,000 -r O O 05 ^ (0 3 0 0.2 0.4 0.6 0.8 1 FIsfiing mortality (u) Figure 3 Optimal reserve proportions and corresponding yields. (A) Panulirus penicillatus. Red Sea spiny lob- ster. [B) Balistes vetula, queen triggerfish. (C ) Haemulon ptumieri, white grunt. iTO^ Epinephelusguttatus, red hind. In all graphs, the solid circles and line represent the sustainable yield (kg of catch per year from the whole management area) that occurs in the absence of a reserve, the open circles and dotted line represent the optimal reser\'e proportion (that which produced maximum sustainable yields for each fishing mortality), and the dashed line and squares represent the sustainable yield when the optimal reserve proportion was used. Intrinsic population growth rates (X) determine the robustness of the populations to fishing, high growth rates sustaining heavy fishing and low rates requiring reserves at low fishing mortalities. This measure is better than standard deviation alone which would treat a 10-kg fluctuation equally, re- gardless of whether it occurred in a 100 kg or 1,000,000 kg per year fishery. We graphed these re- sults with reserve proportion as the independent variate and examined the graphs for trends. We performed these analyses on four coral reef fish- ery species for which we obtained relatively complete parameter sets. These included Balistes vetula, queen triggerfish; Epinepheliis gitttatus, red hind; Hae- mulon plumteri, white grunt; and Panulirus penicil- latus, Red Sea spiny lobster (see Table 1 for param- eter estimates). Results When we ran the models without a reserve (s=0), they produced standard yield-effort curves (Fig. 3). These curves are characterized by steep initial gains in long- term sustainable yields with increases in fishing mortality (and thus effort), followed by equally steep declines (Clark, 1990). The curves peaked at the maximum sustainable yield, one of several goals a manager might try to achieve with a fishery (Clark, 1990), and we will refer to the corresponding fishing mortality as the MSY mortality for the rest of this paper. Above the MSY mortality, the fishery can be defined as overfished because it is less productive than it would be with less fishing activity. When a reserve was present, the yield-mortality curves were still parabolas passing through the ori- gin but spread farther to the right, and the larger the reserve, the more pronounced were these shifts. Consequently, larger reserves required higher fish- ing mortalities to maximize long-term sustainable yields (remember that this mortality only affected fish in fishing areas), whereas the sustainable yields 610 Fishery Bulletin 97(3), 1999 decreased more slowly as fishing mortality increased past the MSY mortality. Our analyses of optimal reserve proportions pro- duced several key results. First, reserves produced fisheries enhancements, meaning that the overall catches with a reserve exceeded those without one, whenever the fisheries were overfished (Fig. 31, here defined as fished above the MSY mortality level. When fisheries were overfished, they produced higher yields with a reserve even though the reserve decreased the amount of fishing area. The optimal reserve pro- portion increased with increasing fishing mortality, and heavily exploited fisheries required particularly large reserves to remain productive. The fishery benefit at- tributable to reserves, calculated by subtracting the yield without a reserve ft-om that with an optimally sized reserve, increased with increasing fishing mor- tality up to a near-maximum yield in most cases ( Fig. 4). Consequently, a wide span of reserve sizes (up to 80% of the management area for some species) pro- 1,250- A Panulirus penicillatus 1 ,250 - B Balisles vetula 1,500- 0.2 0.4 0.6 0.8 1 C Haemulon plumien 0.2 0.4 0.6 0.8 1 1 2,000 T D Epinephelus guttau 0 0.2 0.4 0.6 0.8 1 0 0.2 0,4 0.6 0,8 1 Fishing mortality (u) Figure 4 Catch enhancements with the u.se of an optimally proportioned reserve (OPRl. (A) Panulirus penicillatux. Red Sea spiny lobster. (B) Balistes vetula, queen triggerfish. (C) Haemulon plumieri, white grunt. (D) Epinephelus guttatus. red hind. Value.s represent the increase in yield, in kg of catch per year from the whole management area, one could expect if an optimally sized reserve system were established in a man- agement area that lacked reserves initially. duced similarly high yields for most species as long as fishing mortalities were chosen accordingly. Using this information (Fig. 3), we predicted opti- mal reserve proportions under real-life fishing mor- talities. For queen triggerfish, the fishing mortality estimate of « = 0.45 from Puerto Rico and the Virgin Islands (Aiken, 1983) corresponded to an optimal reserve proportion of approximately s = 0.8. For white grunt, a reported heavy fishing mortality ofu = 0.99 from Jamaica (Darcy, 1983) corresponded to an opti- mal reserve proportion of just over s - 0.75. Thus, for these species in these locations, our models pre- dicted that 75-80% of the fishing grounds should be made off-limits to fishing in order to maximize long- term sustainable yields. These numbers may seem unrealistically high, especially since most models predict maximum yields when approximately 50% of the population density at carrying capacity is pro- tected from fishing ( see Clark, 1990, for an overview ). In the case of our models, populations within the reserve did not reach carrying capacity when fishing was heavy outside, and the conditions of peak production corre- sponded to those that protected approxi- mately 50% of the population density at carrying capacity. The qualitative conclusions outlined above were consistent across all the spe- cies we examined. However, the model's quantitative predictions of the long-term fishery yields and optimal reserve propor- tion varied from species to species for any given fishing mortality (Fig. 3). The key differences between species were the speeds at which the yield and optimal re- serve proportion changed with increasing fishing mortality ( Fig. 3 ). These differences reflected differences in intrinsic population growth rates (A) — the maximum growth rate of a population with no density-depen- dent constraints or fishing mortality. This summary parameter integrates most of the life history data that we used. It does not include the growth rate of individuals in the population and consequently does not adequately predict yields. However, it is a useful summary of the ability of a popula- tion to sustain harvesting. For example, life history parameters from the literature suggested that the Red Sea spiny lobster had a relatively low A = 1.08, just above the A = 1 necessary for a population to sus- tain itself with no fishing pressure. This species had a low MSY fishing mortality because its slow population growth could Sladek Nowlis and Roberts: Fisheries benefits and optimal design of marine reserves 61 only sustain modest harvesting ef- fort (Fig. 31. In contrast, life history parameters from the literature sug- gested that red hind had a relatively high A = 1.31. Consequently, its maximum sustainable yield oc- curred at the highest fishing mor- tality of any species we tested (Fig. 3). The two other species we exam- ined had intermediate intrinsic rates of population growth rates and responses to reserves. The sensitivity of our models' quantitative predictions was also clear within a species when we var- ied larval survivorship. For all spe- cies, optimal reserve proportion and yield without a reserve varied greatly (senior author's unpubl. data) because we varied larval sur- vivorship from 10"^ to 10"*. This sensitivity to poorly understood pa- rameter values renders any quan- titative estimates of optimal reserve proportion unreliable, whether the inaccuracy is in larval sui-vivorship, the relationship, or parameters for density dependence, or any other life history parameter. Finally, we examined how re- serves might influence unpredict- able catches resulting from environ- mental variation. Our stochastic models predicted that catches will be more stable with larger resei-ve proportions. In these models, we saw general decreases in catch variability with in- creasing reserve proportion (Fig. 5). The results pre- sented here showed drops in variation that were more pronounced at higher fishing mortalities for all four species. We also tested these results at three levels of environmental variation. Our results showed that the drop in catch variability was most extreme when the environment was most variable, suggesting that the stability offered by reserves will be most valu- able in highly variable fisheries. Discussion Effects of life history and fishing mortality on reserve benefits Our models predicted that marine fishery reserves will provide catch enhancements to any overfished fishery that meets our basic assumptions regarding Fishing 1 mortalities 0 4 ••»■ 06 ■•- ■ o---- 0 8 -- o- - Reserve proportion (s) Figure 5 Catch variability and reserve size. (A) Panulirus penicillatus. Red Sea spiny lobster. (S) Balistes vetula. the queen triggerfish. IC) Haemulon plumien, white grunt. iD) Epinephelus guttatus, red hind. Each graph shows decreasing catch variability with increasing reserve prnportion at four levels of fishing mortality. the movement of adults and larvae. The results from previous modeling efforts by Man and colleagues (1995) and Holland and co-workers (Holland and Brazee, 1996; Holland et al.') support these findings if one compares their results in specific cases to the patterns we found for a variety of species. Two key variables help determine whether a population is overfished: intrinsic population growth rate (A) and fishing mortality. Managers can control fishing mor- tality to varying extents. Apparently, this control is inadequate in many industrial fisheries (FAO, 1995) and is probably even less effective in subsistence fish- eries (Roberts and Polunin, 1993). Managers have no control over population growth potential but can take into account that species with low population growth have a greater tendency to be overfished and consequently show greater promise for fisheries en- hancements from reserves. Even in a well-managed fishery, it may be helpful to close large areas. This strategy could allow the relaxation of some fishing restrictions in remaining waters. Consequently, recreational and commercial 612 Fishery Bulletin 97(3), 1999 fishermen may feel greater equity with fewer restric- tions on the number of participants or their catches. Moreover, reserves have the potential to reduce vari- ability in catches from year to year and to enhance conservation of species and ecosystems. Fishing is not the only threat to marine ecosystems, though, and fisheries regulations are not sufficient to pro- tect these systems (Allison et al., 1998). To our knowledge, no field study has yet examined the effects of population growth potential or fishing mortality on reserve benefits. In part, such studies are made difficult by the uncontrolled nature in which reserves are established. Relation of fishery benefits to reserve size Real-world fisheries span a range from lightly fished to heavily overfished, and the optimal reserve size will depend on the fishing mortality as well as the population growth potential of the target species. Because many fisheries involve multiple species with widely divergent population growth potentials, choos- ing a single best reserve size may be difficult. More- over, key aspects of the life history of marine fish, the larval phase in particular, remain a mystery. Be- cause of these gaps in knowledge, it would be difficult to make an accurate prediction of the optimal reserve size even in a well-studied single-species fishery. Although our research sheds doubt on the use of a universal reserve proportion, it does lend support for the use of large reserve systems under certain cir- cumstances. In the two real-world cases where the necessary information existed, our models predicted that reserves should encompass 75-809^ of the man- agement area. These proportions are enormous and may be unrealistic for several reasons. First, the short-term economic losses from closing 80'7c of a management area would be large, although our mod- els predict that the recovery time for such heavily overfished fisheries would be rapid (Sladek Nowlis and Roberts, 1997). Second, the political challenges of establishing such large reserves would be a formi- dable barrier Finally, we do not stand firmly behind these predictions because their accuracy is depen- dent on parameter values that are poorly understood. Nevertheless, consistent results across several spe- cies suggest that reserves encompassing 40'7( or more of a heavily fished management area could produce substantial fisheries benefits. Though rare, at least one large reserve system does exist. The Mombasa Marine National Park closed over 60'7c of local fishing grounds (McClanahan and Kaunda-Arara, 1996). This example fits nicely with our model's assumptions because levels of fishing effort remained similar in the fishing grounds be- fore and after the closure. After two years, total yields had not surpassed those prior to reserve establish- ment (McClanahan and Kaunda-Arara, 1996). How- ever, catch per unit of effort had increased dramati- cally and total yields showed potential for future in- creases. In this case and others involving extensive use of marine reserves, our research encourages an adaptive approach that reflects the lack of knowl- edge about fish life histories and the high degree of uncertainty in these complex biological systems. Relation of reserve size to catch variability Our model supported Bohnsack's (1996) hypothesis that catch variability will decrease with increasing reserve size. Our models predicted decreases in catch variability across a variety of levels of environmen- tal variability and fishing mortalities. Our results also complement other studies that showed that re- serves could reduce catch variability,'^ decrease the likelihood of bad years (Lauck et al., 1998), and in- crease the persistence of fisheries vulnerable to over- fishing (senior author's unpubl. data). To our knowledge, no field study has yet examined the effects of reserves on catch variability. Although they may be confounded by variability in fishing ef- fort, the necessary data should be practical to collect before and after reserve creation. Assumptions revisited As with all models, one must be careful in interpret- ing the results of this one. It is based on parameter values that in some incidences — larval survivorship in particular — are poorly understood. However, the model's predictions are qualitatively robust to param- eter errors, meaning that its general predictions hold true across a wide range of values and a wide vari- ety of species. Our assumptions regarding the move- ment of adults and larvae were far more critical in influencing the conclusions we have drawn here. Our assumptions regarding adult movement have wide applicability. Many fisheries target sessile or- ganisms such as harvested kelp (Bustamente and Castilla, 1990), slow-moving organisms including many invertebrates (Davis and Dodrill, 1980; Davis and Dodrill, 1989), and organisms with high site- specificity such as many reef fish (Polunin and Rob- erts, 1996). All of these systems are likely to approxi- mate our assumptions of no adult movement. This model is not universally applicable, as highly mobile and migratory species, including many pelagic fisher- ^ Mangel, M 1998. Environmental Studies Board, University of California, Santa Cruz, CA 95064. Unpubl. data. Sladek Nowlis and Roberts; Fisheries benefits and optimal design of marine reserves 613 ies (Safina, 1993 ), will only fit our adult movement as- sumptions if large reserves are established. Recent studies, though, have shown high site fidelity by fish species previously thought to range widely (Holland et al., 1993; Holland et al., 1996), demonstrating the need for more field data on adult movement patterns. More- over, recent modeling efforts by Holland and colleagues' and others'^ suggest that reserves can benefit highly mobile species through enhanced population fecundity gained fi-om temporary protection. If adults do cross reserve boundaries, our predic- tions regarding fisheries benefits from reserves will be influenced in opposing ways. Under heavy fish- ing pressure and intermediate movement tendencies, minor yield enhancements may be possible from this adult spillover (Polacheck, 1990; DeMartini, 1993). However, this same movement would dilute the abil- ity of reserves to enhance larval transport to fishing areas. As Polacheck showed (1990), spawning stock biomass, or the potential for fisheries enhancement through larval transport, is highest at lowest levels of adult movement. Because the potential benefits from larval transport presented here far outweigh those predicted from adult spillover (Polacheck, 1990; DeMartini, 1993), it is likely that adult movement across boundaries will decrease the predicted yields from reserves. Consequently, reserves will have the highest potential for enhancing surrounding fisher- ies if they are designed as a collection of units large enough to contain populations of adults with rela- tively little movement across boundaries. Our assumptions regarding larval transport have less supporting evidence. Most aquatic species dis- perse more widely as larvae than as adults (Boehlert, 1996), and the potential for long-distance dispersal across reserve boundaries is great for species with long-lived larvae ( Roberts, 1997 ), including most food fish. Consequently, lai-vae are likely to move from reserves to fishing areas as long as oceanographic conditions and larval behavior permit. Without lar- val transport, the potential for fisheries benefits fi'om reserves is more limited, although Holland and col- leagues' did show that a reser\'e system in which lar- vae stayed in place but adults moved widely across boundaries could produce some benefits. Reser\'es, es- pecially in heavily overfished or large management areas, may need to be partitioned into several subunits that maintain adult populations within them but al- low larvae to disperse to remaining fishing areas. We also assumed a stock-recruitment relationship, implying that a significant portion of the population ^ Guenette, S. 1998. Fisheries Centre, University of British Columbia, 2204 Main Mall, Vancouver. BC V6T 1Z4. Canada. Unpublished data. fecundity from reserves stays in or returns to the management area. The degree to which marine popu- lations are locally sustained remains an active area of debate in marine ecology. Larvae of most tropical food fish are often found in greatest quantities off- shore (Boehlert, 1996), suggesting the possibility of long-distance dispersal. However, studies that show this result may be biased because sampling within the complex structure of the reef itself is difficult ( Boehlert, 1996 ). Therefore, reefs may harbor greater concentrations of larvae than are measured above the reef This complexity (Wolanski and Sarsenski, 1997), along with potential for larval behavior to in- fluence their distribution (e.g. Breitburg et al., 1995), suggests that larvae may be retained at higher con- centrations than predicted by simple oceanographic models (e.g. Roberts, 1997). If recruitment dynam- ics are influenced on a much larger spatial scale than encompassed by the management area, such that the stock in the management area has a minimal im- pact on recruitment back to it, reser\'e benefits to the management area are likely to be much more limited. Cohort models, including those by Polacheck (1990) and DeMartini (1993), can be interpreted as situations in which larval supply is constant and not influenced by local stock. As has been discussed, these models show limited potential for fisheries benefits from reser\'es. It is necessary to think of reserve sys- tems at a scale that fits stock-recruitment relation- ships. Yet our knowledge of these relationships re- mains poor. Even if larvae have the potential to dis- perse over large distances, stock-recruitment rela- tionships could still exist on a local level if a signifi- cant portion of larval production is retained. The safest approach to this uncertainty is to design re- serve systems at large scales. However, there is still the potential for reserves to produce fisheries ben- efits on small scales if lai-vae have the capacity to be retained. Further research on stock-recruitment re- lationships in marine populations will be invaluable for resolving this pressing issue along with many others in fisheries management. Field needs and testable predictions Our results identify areas in need of additional field work and make testable predictions. The needs in regard to field work differ for our quantitative and qualitative predictions. The quantitative predictions were highly sensitive to all parameters that affected intrinsic population growth potential. The most im- portant and least understood of these parameters is lai'val sui-v'ivorship. We need significantly better in- formation about the duration of the egg and larval stages of coral reef fishes and their daily mortality 614 Fishery Bulletin 97(3), 1999 risk. Until we understand these life history stages better, it will be impossible to make quantitatively accurate predictions of the optimal design of any fish- ery management strategy. We also need better insight into how fecundity changes with size. Fecundity-size relationships should be fairly easy to measure and can be incorpo- rated into any standard fishery study where adequate numbers of adults are sampled. We would further benefit from estimates of size-specific natural mor- tality. Few natural mortality estimates for coral reef fish species exist in the literature, and most that do are based on highly indirect methods of association. Marine fishery reserves actually offer the potential to generate more accurate predictions of natural mortality because fishing mortality does not confound the attempt in unfished areas. Moreover, despite numerous studies, we still have a poor understand- ing of population regulation and density dependence in coral reef fishes. This understanding is also nec- essary before we can generate accurate quantitative predictions of reserve benefits. In contrast to the long list necessary to generate quantitative predictions, our qualitative predictions require additional knowledge in only one key area: fish movement. Because the qualitative predictions were robust across life history patterns, the key to knowing whether a fish species fits our assumptions is the movement of this species as eggs, larvae, and as adults. To some extent, we can skirt this issue because in our model, reserve size was based on pro- portion of coastline rather than actual size. Conse- quently, if we choose the management area to match the scale of fish movement, our model can fit most species. For example, a 20% reserve divided into ar- eas of tens of hectares might ensure that adults of the species we examined here will stay in the area in which they settled while their larvae disperse widely among the reserve and nonreserve areas. In contrast, the management area might have to encompass whole ocean basins for the movement assumptions to fit bluefin tuna (Safina, 1993). Thus, we need to understand the movement dynamics of larvae and adults of a species to know the scales at which it will fit the assumptions of our model. From the species that we ran and the resulting qualitative predictions of our model, we can gener- ate a list of testable predictions. We predict that 1) Reserves will be beneficial for any over- fished population. Populations with low intrinsic growth rates and high fishing mortality stand to benefit the most, as is the case for the majority of reef fisheries in many regions of the world, such as the Caribbean. The location and size of the re- serve will also affect reserve benefits. For a fair test of this prediction, reserve should be repre- sentative of typical fish habitat and large enough to contain a viable population of adults. 2) Although no universal best reserve proportion exists, we predict reserves will enhance fishery pro- ductivity even when they encompass areas much larger than those of current reserve systems. 3 ) Reserves will reduce variation in catches result- ing from unpredictability in fishing mortality as well as recruitment strength and larval survivor- ship. Such an effect will simplify fishery manage- ment and increase the ability of fishermen to pre- dict future income. Acknowledgments We gratefully thank the U. S. Agency for Interna- tional Development and its Research Program for Historically Black Colleges and Universities as well as the University of Puerto Rico Sea Grant College Program and the National Research Council's Re- search Associate Program for generous support of the research reported here. We also extend our gratitude to Hilconida Calumpong of the Silliman University Marine Laboratory, who provided inspiration for this work, and to Rebecca Sladek Nowlis, who provided valuable suggestions on drafts of the manuscript. Literature cited Aiken, K. A. 1983. The biology, ecology and bionomics of the trigger- fishes, Balistidae. In J. L. Munro led.). Caribbean coral reef fishery resources, p. 191-276. International Council for Living Aquatic Resources Management, Manila. Alcala, A. C, and G. R. Russ. 1990. A direct test of the effects of protective management on abundance and yield of tropical marine resources. J. Cons. Int. Explor Mer 47:40-47. Allison, G. W., J. Lubchenco, and M. H. Carr. 1998. Marine reserves are necessary but not sufficient for marine conservation. Ecol. Appl. 8:S79-92. Beverton, R. J. H., and S. J. Holt. 1957. On the dynamics of exploited fish populations. Chapman and Hall, New York. NY. 533 p. Boehlert, G. W. 1996 Larval dispersal and survival in tropical reef fishes. In N. V. C. Polunin and C. M. Roberts (eds.l. Reef fisher- ies, p. 61-84. Chapman and Hall, London. Bohnsack, J. A. 1996 Maintenance and recovery of fishery productivity In N. V. C. Polunin and C. M. Roberts (eds.). Reef fisheries, p. 283-313. Chapman and Hall, London. Breitburg, D. L., M. A. Palmer, and T. Loher. 1995. Larval distributions and the spatial patterns of settle- ment of an oyster reef fish: responses to flow and structure. Mar Ecol. Prog. Sen 125:45-60. Sladek Nowlis and Roberts: Fisheries benefits and optimal design of marine reserves 615 Bustamente, R. H., and J. C. Castilla. 1990. Impact of human exploitation of populations of the intertidal bull-kelp Durcillaea antarctica (Phaeophyta, Durvilleales) in central Chile. Biol. Cons. 52:205-220. Castilla, J. C, and L. R. Duran. 1985. Human exclusion from the rocky intertidal zone of central Chile: the effects on Concholepas concholepas (Gastropoda). Oikos 45:391-399. Clark, C. W. 1990. Mathematical bioeconomics: the optimal manage- ment of renewable resources. John Wiley & Sons, New York. NY, 399 p. Darcy, G. H. 1983. Synopsis of biological data on the grunts Haemulon aurolineatum and H. plumien iPisces: Haemulidae). U.S. Dep. Commer., NOAA Technical Report NMFS Circular 448, 41 p. Davis, G. E., and J. W. Dodrill. 1980. Marine parks and sanctuaries for spiny lobster fish- eries management. Fish. Bull. 78:979-984. 1989. Recreational fishery and population dynamics of spiny lobster, Panulirus argus. in Florida Bay, Everglades National Park, 1977-1980. Bull. Mar Sci. 44:78-88. DeMartini, E. D. 1993. Modeling the potential of fishery reserves for man- aging Pacific coral reef fishes. Fish. Bull. 91:414-427. Dugan, J. E., and G. E. Davis. 1993. Applications of marine refugia to coastal fisheries management. Can. J. Fish. Aquat. Sci. 50:2029-2042. Duran, L. R., and J. C. Castilla. 1989. Variation and persistence of the middle rocky inter- tidal community of central Chile with and without human harvesting. Mar Biol. 103:55.5-562. FAO (Food and Agriculture Organization). 1995. The state of the world fisheries and aquaculture. Food and Agriculture Organization Fisheries Department FAO Fisheries Circular 884, 105 p. Hay, M. E. 1984. Patterns offish and urchin grazing on Caribbean coral reefs: are previous results typical? Ecology 65:446-454. Hay, M. E., and P. R. Taylor. 1985. Competition between herbivorous fishes and urchins on Caribbean reefs. Oecologia 65:591-598. Hixon, M. A. 1991. Predation as a process structuring coral reef fish communities. In P. F Sale (ed.). The ecology of fishes on coral reefs, p. 475-508. Academic Press, San Diego, CA. Hixon, M. A., and J. P. Beets. 1993. Predation. prey refuges, and the structure of coral- reef fish assemblages. Ecol. Mono. 63:77-101. Hixon, M. A., and M. H. Carr. 1997. Synergistic predation, density dependence, and popu- lation regulation in marine fish. Science (Wash. D. C.) 277:946-949. Holland, D. S., and R. J. Brazee. 1996. Marine reserves for fisheries management. Mar Resource Econ. 11:157-171, Holland, K. N., C. G. Lowe, and B. M. Wetherbee 1996. Movements and dispersal patterns of blue trevally iCaranx melampygus) in a fisheries conser\'ation zone. Fish. Res. 25:279-292. Holland, K. N., J. D. Peterson, C. G. Lowe, and B. M. Wetherbee. 1993. Movements, distribution and growth rates of the white goatfish Mulloides flavolitieatus in a fisheries con- servation zone. Bull. Mar Sci. 52:982-992. Houde, E. D. 1989. Comparative growth, mortality, and energetics of marine fish larvae: temperature and implied latitudinal effects. Fish. Bull. 87:471-495. Lauck, T., C. W. Clark, M. Mangel, and G. R. Munro. 1998. Implementing the precautionary principle in fisher- ies management through marine reserves. Ecol. App. 8:S72-78. Man, A., R. Law, and N. V. C. Polunin. 1995. Role of marine reserves in recruitment to reef fisher- ies: a metapopulation model. Biol. Cons. 71:197-204. McAllister, D. E. 1988. Environmental, economic and social costs of coral reef destruction in the Philippines. Galaxea 7:161-178. McClanahan, T. R., A. T. Kamukuru, N. A. Muthiga, M. Gilagabher Yebio, and D. Obura. 1996. Effect of sea urchin reductions on algae, coral, and fish populations. Cons. Biol. 10:136-154. McClanahan, T. R., and B. Kaunda-Arara. 1996. Fishery recovery in a coral-reef marine park and its effect on the adjacent fishery Cons. Biol. 10:1187-1199. McClanahan, T. R., and S. H. Shafir. 1990. Causes and consequences of sea urchin abundance and diversity in Kenyan coral reef lagoons. Oecologia 83:362-370. Pauly, D., V. Christensen, J. Dalsgaard, R. Froese, and F. J. Torres. 1998. Fishing down marine food webs. Science (Wash. D.C.) 279:860-863. Plan Development Team. 1990. The potential of marine fishery reserves for reef fish management in the U.S. Southern Atlantic. U.S. Dep. Commer .Technical Memorandum NMFS-SEFC-261, 48 p. Plant, L 1993. Sexual maturity, reproductive season and fecundity of the spiny lohsteT Panulirus penicillatus from the Gulf of Eilat ( Aqaba l. Red Sea. Aust. J. Mar Freshwater Res. 44:527-535. Plant, I., and L. Fishelson. 1991. Population structure and growth in captivity of the spiny lobster Panulirus penicillatus from Dahab, Gulf of Aqaba, Red Sea. Mar Biol. 111:467-472. Polacheck, T. 1990. Year around closed areas as a management tool. Nat. Resource Model. 4:327-354. Polunin, N. V. C, and C. M. Roberts (eds.). 1996. Reef fisheries. Chapman and Hall, New York, NY, 495 p. Quinn, J. F., S. R. Wing, and L. W. Botsford. 1993. Harvest refugia in marine invertebrate fisheries: models and applications to the red sea urchin Strongylo- centrotus franciscanus. Am. Zool. 33:537-550. Ricker, W. E. 1975. Computation and interpretation of biological statis- tics of fish populations. Fisheries and Marine Service, Ottowa, 400 p. Roberts, C. M. 1995. Effects of fishing on the ecosystem structure of coral reefs. Cons. Biol. 9:988-995. 1997, Connectivity and management of Caribbean coral reefs. Science (Wash, DC.) 278:1454-1457. Roberts, C. M„ W. J. Ballantine, C. D. Buxton, P. Dayton, L. B. Crowdcr, W. Milon, M. K. Orbach, D. Pauly, and J. Trexler. 1995. Review of the use of marine fishery reserves in the U.S. southeastern Atlantic. U.S. Dep. Commer, NOAA Technical Memorandum NMFS-SEFC-376, 31 p. 616 Fishery Bulletin 97(3), 1999 Roberts, C. M., and N. V. C. Polunin. 1991. Are marine reserves effective in management of reef fisheries? Rev Fish Biol. Fish. 1:65-91. 1993. Marine reserves: simple solutions to managing com- plex fisheries? Ambio 22:363-368. Rowley, R. J. 1994. Marine reserves in fisheries management. Aquat. Cons. Mar. Freshwater Ecosyst. 4:233-254. Russ. G. 1985. Effects of protective management on coral reef fishes in the central Philippines. Proceedings of the 5th Inter- national Coral Reef Congress 4:219-224, Russ, G. R., and A. C. Alcala. 1996. Do marine reserves export adult fish biomass? Evi- dence from Apo Island, central Philippines. Mar. Ecol. Prog. Ser. 132:1-9. Safina, C. 1993. Bluefin tuna in the West Atlantic: negligent manage- ment and the making of an endangered species. Cons. Biol. 7:229-234. Sladek Nowlis, J., C. and M. Roberts. 1997. You can have your fish and eat it, too: theoretical approaches to marine reserve design. Proceedings of the 8th International Coral Reef Symposium 2:1907-1910, Smithsonian Tropical Research Institute, Panama City, Panama. Thompson, R., and J. L. Munro. 1983. The biology, ecology and bionomics of the hinds and groupers, Serranidae. /;; J. L. Munro led.), Caribbean coral reef fishery resources, p. 59-81. International Coun- cil for Living Aquatic Resources Management, Manila. Wolanski, E., and J. Sarsenski. 1997. Larvae dispersion in coral reefs and mangroves. Am. Sci. 85:236-243. Appendix Fecundities Fecundities were size-specific, but the general form of the equation relating size to fecundity varied from species to species. The specific relationships are listed in Table 1 as /•. These r values were set to zero for all classes smaller than the size at maturity. Larval survival We used equations developed by Houde ( 1989) that relate ambient temperature during development to duration of larval stage, daily mortality risk, and prob- ability of surviving through the entire larval stage. D = 952.5 7^ 1 .07.52 Z = 0.0003149 7 (11 (2) (3) where T = ambient temperature during develop- ment, in degj-ees Celsius; D = duration of larval stage, in days; Z = probability of mortality, per day; and A'^ = probability of surviving through the en- tire larval stage. Adult survival We assumed that newly settled fish experienced den- sity dependence. Thus, instead of surviving at a rate L'j like individuals in other size classes, their survival was weighted by a density-dependent function of the form e '' '^ where p = the population density and K = a measure of carrying capacity arbitrarily set at 1000 due to a lack of information on carrying capacities for the fish we studied. Note that size-class- 1 indi- viduals included new recruits that survived and grew as well as old size-class-1 individuals that survived but did not grow to size class 2. Thus, at time t, the densities of size-class-1 individuals in the reserve iSj ^) and the fishing area (F^ ^) are Si, = V(,piO}Sof_-^e~^ F,j=VopiO}Foj_,e~ + ri(l-p(l))Si,,_i (4) ,/A- + v,(l- P<1>)^1,M. (5) where v = the density-independent survival rate for individuals in size class .v. Note that the density in the fishing area is de- creased later in the program to account for fishing mortality but only for size classes larger than the size at fisheiy recioiitment. Also note that other size classes experience the density-independent survival rate v^. Growth We began with standard von Bertalanffy equations (Ricker, 1975 1, relating length to age and weight to length (Fig. 2) and categorized them as described by Figure 2. Through algebraic manipulation, we estab- lished a formula for g(B^J, the size of an individual projected one year in the future: ^(Bj^e-^'S +(l-e-''')L,„^- (6) We used this formula to establish the following cal- culation forp(.x'), the probability that an individual in size class x grows to size class x+l by next year. pix) S. B ill. (7) 617 Abstract.— Snow crab iChionoecetes opilio^ and Tanner crab (C bairdi) fish- eries in the eastern Bering Sea are managed by using swept-area esti- mates of biomass based on an annual survey conducted with a 83-112 east- ern bottom trawl. Estimates of net ef- ficiency (i.e. the capture probability of crab that occur between the wing-tips of the trawl net) are needed to correct the biomass estimates for any size se- lectivity by the trawl. Data on net effi- ciency were obtained experimentally by attaching an auxiliary net beneath the trawl net to capture crab escaping un- der the trawl footrope. Net efficiency is then the quotient of the trawl catch di- vided by the combined catch of the trawl and auxiliary nets. Mathemati- cal models of the relationship between net efficiency and carapace width were formulated and fitted to the experimen- tal data. Net efficiency for both species first decreased with increasing cara- pace width until a minimum efficiency was reached near 50 mm carapace width. At larger sizes, efficiency in- creased asymptotically with carapace width. Net efficiency for mature female Tanner crab was lower 10.47) than for males and immature females combined (0.72) at the same mean carapace width as mature females (66-107 mm). Net efficiency did not differ between mor- phologically mature and immature male Tanner crabs of the same carapace width. Net efficiency of a survey trawl for snow crab, Chionoecetes opilio, and Tanner crab, C bairdi David A. Somerton Alaska Fisheries Science Center National Marine Fisheries Service, NOAA 7600 Sand Point Way NE Seattle, Washington 98125 E-mail address david somertonanoaa gov Robert S. Otto Kodiak Laboratory Alaska Fisheries Science Center National Marine Fishenes Service, NOAA PO Box 1638 Kodiak, Alaska 99615 Manuscript accepted 3 November 1998. Fish. Bull. 97:617-625 ( 1999). The commercial fisheries for snow crab (Chionoecetes opilio) and Tan- ner crab (C. bairdi) in the eastern Bering Sea are managed by using swept-area estimates of biomass (Alverson and Pereyra, 1969) pro- vided by an annual Alaska Fisher- ies Science Center (AFSC) bottom trawl survey. In addition to catch and swept-area data, such biomass estimates require a value for the sampling efficiency of the trawP (i.e. the proportion of animals that are captured within the area spanned by the trawl doors; Dickson, 1993a). Trawl efficiency can be considered as a function of sweep efficiency (the proportion of animals within the path of the doors, bridles, and sweeps that are herded into the net path) and net efficiency (the propor- tion of animals that are captured within the path of the trawl net; Dickson, 1993a). For snow and Tan- ner crabs, owing to a lack of any contradictory evidence, sweep effi- ciency was assumed to be zero and net efficiency was assumed to be unit}'. The assumption of zero sweep efficiency has been subsequently supported by experiments on herd- ing,^ but the assumption of com- plete net efficiency has been contra- dicted by studies in which low-light video cameras have documented that even large male crabs could often escape under the footrope.'^ Such observations prompted us to experimentally estimate net effi- ciency of the 83-112 eastern trawl, the trawl used for the eastern Bering Sea surveys. Previous studies of net efficiency, which were focused at various spe- cies of groundfish, indicated that efficiency generally increases with body size (Engas and God0, 1989; Walsh, 1992). For snow and Tanner crab, however, biological attributes other than size could also influence efficiency. For example, mature males have longer legs and thicker chelae, in relation to their carapace width, than either mature females or immature males. Not only could the difference in body shape result in differing net efficiency, differ- ' Throughout this paper we will refer to the trawl as an entire fishing gear comprising the net. bridles, and doors. - Somerton, D. A., and P. T Munro. 1998. Estimating the sweep efficiency of a bottom trawl. Manuscript in preparation. ' Munro, P. T. 1998. Alaska Fisheries Sci- ence Center, 7600 Sand Point Way NE Se- attle, WA. Unpubl. data. 618 Fishery Bulletin 97(3), 1999 I ences in behavior between sexes or between mature and immature indi- viduals but could also lead to differ- ences in efficiency. Of particular con- cern is a possible difference in net efficiency between mature and imma- ture males because such a difference could lead to a bias in the estimates of the carapace width at maturity and thereby affect commercial mini- mum width regulations. We estimated net efficiency with an experimental approach pioneered by Engas and Godo ( 1989) that uses an auxiliary net attached beneath the trawl net to capture crab escaping under the footrope. If the auxiliary net spans the fishing width of the trawl net, from wing-tip to wing-tip, and captures everything passing be- neath the footrope, then efficiency is estimated as the quotient of the trawl catch divided by the total catch of both the trawl and auxiliary net (Walsh, 1992; Dickson 1993b). The crux of such an experiment is then to design an auxiliary net with a footrope that exerts sufficient bottom contact to capture everything pass- ing beneath the trawl footrope with- out causing enough added drag to al- ter the fishing efficiency of the trawl. Unlike the trawls examined in the studies of Engas and God0( 1989) and Walsh (1992), the 83-112 trawl lacks bobbins or rollers on the footrope and instead has a footrope consisting of a simple, rubber-wrapped cable. This difference in trawl design required several distinct differences in the design of the auxiliary net. In this paper we examine the application of this auxiliary net to estimate the efficiency of the 83-112 eastern trawl for snow and Tanner crabs. Materials and methods Description of the trawl and auxiliary net The 83-112 eastern trawl is a two-seam trawl with a 25.3-m headrope and a 34.1-m fpotrope consisting of a simple rubber-wrapped cable intended to remain in close contact with a smooth bottom (Fig. 1; see Armistead and Nichol, 1993, for a detailed net plan). The trawl is towed behind 1.8 x 2.7 m steel "V" doors Auxiliary net Auxiliary net footrope Figure 1 The 83-112 eastern trawl net and the auxilary net. The dashed lines show the corresponding attachment points on both the trawl and auxilary nets, but the two nets are sewed together along the entire edge of the auxilary net. weighing 816 kg each, which are connected to each wing with two 55-m bridles. Wing and throat sec- tions of the net are made of 10.1-cm (stretched mea- sure) nylon mesh, the intermediate section is made of 8.9-cm mesh and the codend is made with a double layer of 8.9-cm mesh lined with 3.1-cm mesh. The auxiliary net was hung beneath the trawl net, attached at the wing tips and along the top of the wings (Fig. 1). Unlike the auxiliary net designs of Engas and Godo (1989 ) and Walsh ( 1992 ), our auxil- iary net lacked a headrope, used the belly mesh of the trawl as the top panel, and had a single codend rather than three separate codends. The mesh used in the auxiliary net was the same as that for the trawl, but the footrope was longer ( 38.2 m) and was constructed of 16-mm long-link steel chain intended to be sufficiently heavy to excavate buried crab. An Somerton and Otto: Net efficiency of a survey trawl for Ch/onoecetes opilio and C, bairdi 619 additional feature of the auxiliary net was that the footrope was not attached anywhere along the length of the trawl footrope but instead joined it only at the wing tips. The auxiliary net footrope was intended to be positioned 2-3 m behind the trawl footrope along the center line of the trawl, but this position was not verified in the field. eled the relationship between net efficiency and cara- pace width as a function of two width-dependent pro- cesses: the probability of entering the net at the footrope (P,) and the probability of entering the net through the belly mesh after escaping under the footrope (Pj,). Algebraically this relationship can be expressed as Description of the experiment The net efficiency experiment was conducted aboard a chartered 40-m stern trawler (YV Arcturus), from 20 July 1997 to 8 August 1997, immediately follow- ing completion of the annual AFSC trawl survey of the eastern Bering Sea. The experimental site was located east of the Pribilof Islands (57°11.4'N, 166°26.5'W) at depths between 75 and 90 m on a smooth bottom. All hauls were made by using the same trawling procedure and towing speed (3 knots) as those used during the AFSC survey but tow duration ( 15 min) was only one half of the standard. Trawl net width, from wing-tip to wing-tip, was measured on all hauls with a SCANMAR net mensuration system. The experiment began with test hauls of both the standard trawl and the trawl with the attached aux- iliary net (experimental trawl) to determine if trawl performance was altered by the auxiliary net. All test hauls were conducted with the codends left open and with a video camera attached so that the trawl footrope near its center could be viewed. After three hauls of the experimental trawl, it became evident that the additional drag of the auxiliary net resulted in a reduction in the opening width of the trawl net. To help correct this problem, the bridles were short- ened from the standard length of 55 m to 28 m. The shorter bridles were used for the remainder or the experiment. All crab were removed from each catch and sorted into two groups; mature females, which were recog- nized by the presence of an enlarged abdomen, and all other crab. For simplicity, we will refer to the group "all other crab" as "mixed sexes." Carapace width, to the nearest 1 mm, was measured with vernier calipers on all individuals. On some hauls, randomly selected male Tanner crab were subsequently removed from the mixed sexes baskets and measured for both carapace width and chela height (right hand side) to allow de- termination of morphometric maturity. Estimating net efficiency Net efficiency (E^^) typically is an increasing func- tion of body size (Engas and Godo, 1989; Walsh, 1992). To provide more flexibility, however, we mod- E„=Pf^(^-P,^P,- (1) The probability of entering the trawl at the footrope, was described with a three-parameter logistic function: where a, b, and c w 1 + be'-"" parameters; and carapace width. (2) This function was chosen because it provides pre- dicted values of P, that increase with w asymptoti- cally to a value of a and permits cases where effi- ciency is less than unity even at the largest body sizes. The probability of entering the trawl through the belly mesh was described by a two-parameter decreasing logistic function: 1 + rfe'-^"' where d and f = parameters. This function was chosen because it provides pre- dicted values of P/, that decrease with w asymptoti- cally to a value of zero. The model resulting from substituting Equations 2 and 3 into Equation 1 was fitted to values of net efficiency for each 1-mm increment of carapace width by using a maximum likelihood procedure where the number of crab captured in the trawl net at each carapace width was described by a binomial random variable (Millar, 1992). The parameters of the model were estimated by minimizing the negative logarithm of the likelihood function: L = -Y^{n,.\ogE„,,.+(N,,-nJloga- £„.„,)), (4) where, at each value of carapace width w, n = the number of crab captured in the trawl net; TV = the total number of crab in the trawl If and auxiliary nets combined; and E = the net efficiency. 620 Fishery Bulletin 97(3), 1999 A^^^. and n^^ were calculated by pooling together the data from all good tows. For mature female crab, a simplified model consisting of only the P. term was fitted to data because the smallest mature females were too large to fit through the belly meshes of the trawl net. In both cases, the models were fitted to the data by using the S+ function MS (Venables and Ripley, 1994). The 95% confidence intervals for E^^ ^^, were esti- piated by using a bootstrap analysis (Efron and Tibshirani, 1993) which considered between-haul variability but ignored within-haul binomial variabil- ity which is relatively small. Each bootstrap estimate of E^ ^j was calculated by randomly sampling indi- vidual hauls with replacement from the original data, then by fitting the model to the data from the chosen hauls pooled together. After replicating the bootstrap process 240 times, values oiE^^ ^^. at each increment of carapace width were then ranked. The upper and lower confidence intervals were chosen as the values of fi^^ ^^. ranked 7* and 234* at each width increment. Two tests of model form were conducted. First, to determine if the P^ term was a significant contribu- tion to the model, the goodness of fit of the model, including both the P, and P^ terms (five parameters), was compared to that of a model including only the P. term (three parameters) by using a likelihood ra- tio test (pages 153-155 in Hilborn and Mangel, 1997). Second, to determine if the models for mixed sexes differed between crab species, the summed likeli- hoods of the model fit to each species were compared with the likelihood from a model fitted to the com- bined data for both species by using a likelihood ra- tio test. For both tests, significance of the likelihood ratio was evaluated by using a chi square statistic with degrees of freedom equal to the difference in parameters between the models being tested (Hilborn and Mangel, 1997). In addition to the tests on model form, two tests were also conducted for differences in net efficiency between biological subgroups of Tanner crab. First, a test was conducted to determine whether net effi- ciency differs between mature females and mixed sexes restricted to the same range of carapace width as mature females. Because of the restricted range of carapace widths, we assumed that net efficiency could be modeled as a simple logistic function of cara- pace width and sex. This model was fitted by using the S-i- function GLM (Venables and Ripley, 1994). Second, a test was conducted to determine whether net efficiency for mature male Tanner crab differed from that for immature male Tanner crab of equal size. Male crab were categorized as either immature or mature on the basis of height of their chela in re- lation to their carapace width by using the computer technique of Somerton (1980). Both categories were then restricted to a common range of carapace width that was spanned by the smallest mature and the largest immature individuals. Net efficiency was then modeled as a function of maturity and carapace width by using logistic regression. Significance of the sex term in the first test and the maturity term in the second test were assessed with analysis of deviance (page 186 in Venables and Ripley, 1994). When a term was significant, the resulting model was evaluated to predict net efficiency for each biological class at the midpoint of the carapace width interval examined. Results Effect of the auxiliary net on trawl performance The attachment of the auxiliary net to the trawl net resulted in a decrease in wing spread, presumably from increased drag on the bottom. During the test- ing phase when the codends of both the trawl and auxiliary nets remained open, average wing spread for hauls with standard 55-m bridles was 14.3 m (n=3), considerably less than the mean width of the standard net (mean= 17.0, n =29) used during the pre- ceding survey hauls with the same length of towing cable. To help compensate for the additional drag of the auxiliary net, we shortened the bridles to 28 m. This increased average wing spread to 15.4 m (n=4). Wlien the codends were closed, however, the auxil- iary net rapidly filled with epibenthic fauna, espe- cially brittlestars, resulting in a progressive narrow- ing of the wing spread by as much as 3 m over the duration of a haul. After seven attempts at trawling in the initial study area, we were forced to locate a new study area with a lower abundance of brittle- stars. The progressive narrowing was nearly elimi- nated by the change of sampling site, but the mean wing spread during the remainder of the experimen- tal hauls (mean=15.9, ^(=24) was still significantly less (^test, ^=5.18, P<0.001) than the standard net. Although the wing spread was less during the ex- periment than during the AFSC survey, the depar- ture from the standard width may not have been great enough to affect footrope contact with the bot- tom. Footrope contact at the center of the trawl was evaluated by using a video camera for three tows of the standard trawl (mean width=17.5 m), three tows of the experimental trawl with standard bridles (mean width=14.3 m), and three tows of the experi- mental trawl with short bridles (mean width=15.3 m ). In all cases, the footrope in the bosom of the trawl did not contact the bottom, as evidenced by the lack of mud clouds behind the footrope, but instead Somerton and Otto Net efficiency of a survey trawl for Chionoecetes optlto and C bairdi 621 skimmed over the bottom at an estimated height of 1-2 cm. No difference in the proximity of the footrope to the bottom could be visually detected among the three trawl configurations. Be- cause the range in wing spread among the three configurations was greater than the difference between the survey average and the experiment average, it is likely that the addition of the auxiliary net did not alter the contact of the footrope in the cen- ter of the trawl where most of the escapement usually occurs (Walsh, 1992). Description of the data Tanner crab Mixed sexes Snow crab Mixed sexes 200- • • 150 • 100- 50- • n = 3487 Ol^ ISO- IOC 50 0\ n = 3458 ;• A i % • r\ ... 50 100 150 50 100 150 J3 E Mature females 30 20 10 Thirty-one hauls of the experi- mental trawl were completed, but the data from the first seven were not used in subsequent analysis because of unusual trawl performance (the narrow- ing of wing spread associated with large catches of epibenthic fauna in the auxiliary net I. For snow crab, width measurements were collected from 3458 indi- viduals from the mixed sexes cat- egory, but from only three indi- viduals from the mature female category. The low sample size for mature females prevented further analysis. Cara- pace-width frequency distributions indicated that a broad range of carapace widths was sampled, but the observations were concentrated in three carapace width modes (Fig. 2). For Tanner crab, width mea- surements were collected from 3487 individuals from the mixed sexes category, and from 579 individuals from the mature female category. Sampling for both categories covered a broad range of carapace widths, but for the mixed sexes category the observations were concentrated at smaller carapace widths. Plots of net efficiency against carapace width (Fig. 3) for the mixed sexes groups of both species showed an initial decline with increasing width until a mini- mum in net efficiency was reached in the range of 40-50 mm, then a steady increase as carapace width increased. For mature female Tanner crab, the nar- row range of carapace widths and the large variance in net efficiency obscured any obvious change in net efficiency with carapace width (Fig. 3). 50 100 150 Carapace width (mm) Figure 2 Carapace-width frequency distributions for the mixed sexes of snow crab [Chionoecetes opilio) and for the mixed sexes and mature females of Tanner crab iC. bairdi \. The fitted models For the mixed-sexes category of both species, the models that included both the P^and P^ terms fitted the data significantly better than the models with only the P. term (Tanner crab, likelihood ratio= 189.7, P<6.001; snow crab, likelihood ratio=60.5, P<0.001 ). This indicates that the apparent increase in net effi- ciency at small sizes is statistically detectable. Like- wise, the separate-species model fitted the net effi- ciency data significantly better than did the combined species model (likelihood ratio=85.5, P<0.001). This indicates that the change in net efficiency with width differs between species. Plots of the fitted models (Fig. 3) indicated that the predicted values of P^ de- cline rapidly with increasing carapace width and be- come negligible at about 50 mm for Tanner crab and 60 mm for snow crab. For mature female Tanner crab, the model considering only the P.. term fitted the data significantly better than did the sample mean (likeli- 622 Fishery Bulletin 97(3), 1999 hood ratio=4.27, P=0.014), indicating that net efficiency increased significantly over the narrow width range of mature females (66-107 mm). Estimated parameter values for the fitted models are shown in Table 1. For male Tanner crab, net efficiency for mature males (0.82, n-99) was not significantly different (analysis of deviance, P=0.47) than for immature males (0.81, n=39) at the center of the width inter- val spanned by the largest immature and the small- est mature individuals. This indicates that mature males are not preferentially selected by the survey trawl. For mafure female Tanner crab, net efficiency (0.47, n=578) was significantly different (analysis of deviance, P<0.001) than the value for the mixed-sex group (0.72, 77=441) at the midpoint of the mature female size range. This finding indicates that net efficiency was substantially less for mature females than for the mixed-sexes category (combined sexes excluding mature females). Tanner crab Mixed sexes Snow crab Mixed sexes 1.0- • 0.8- % 0.6- ■\. 0.4- - V 1 0.2- -\ 0.0- « 50 100 150 50 Mature females 1.0 0.8- 0.6 0.4- ••••• 0.2 •• 0.0- _ 50 100 150 Carapace width (mm) Figure 3 Values of net efficiency by 1-mm intervals of carapace width (filled circles) and the fitted model (dashed line). Also shown for the mixed sexes categories of both spe- cies are the two model components, P. (solid line ascending to the right) and P^ (solid line decending to the right). Discussion Experimental trawl performance Net efficiency estimates from auxiliary net experi- ments are potentially subject to two sources of bias. First, bias may result if the efficiency of the experi- mental trawl differs from that of the standard trawl. Of particular concern is any distortion of trawl ge- ometry caused by the auxiliary net that changes the position of the footrope and fishing line in relation to the bottom. In our case, trawl net width was reduced by the increased drag of the auxiliary net. We at- tempted to compensate for this increased drag by reducing the bridles to one half of their standard length, but the resulting trawl net width was still approximately one meter less than the standard. The decreased trawl net width, however, apparently did not alter the proximity of the footrope to the bottom because our video observations did not reveal any conspicuous differences between the standard and experimental trawls. Al- though these observations were confined to the center of the footrope, previous video observa- tions of the 83-112 trawl indi- cated that the center of the footrope tends to have lighter bot- tom contact than the wings and is therefore the area most likely to be impacted by the auxiliary net. Second, bias may result if the experimental trawl cannot sample all habitats sampled by the stan- dard trawl. To be effective, the footrope of an auxiliary net must fish harder on the bottom than the footrope of a trawl net. Con- sequently, the auxiliary net is more prone to damage from snag- ging on rocks and from filling with epibenthos and debris. Al- though, in our case, the chain footrope on the auxiliary net may have been effective at excavating buried crab, it also resulted in large catches of epibenthic fauna which, in turn, aggravated the problem of narrowed trawl width. We reduced the narrowing by moving the experiment to a loca- tion lacking high quantities of epibenthos, but by doing so we 100 150 Somerton and Otto: Net efficiency of a survey trawl for Chionoecetes opilio and C. bairdi 623 Table 1 Parameter estimates of the fitted models for the mi.xed | sexes of snow crab iChion oecetes opilio) an d the mixed sexes and mature female categories of Tanner crab (C bairdi). Parameter Snow crab Tanner crab Tanner crab estimates mixed sexes mixed sexes mature females a 0.9894 0.8743 0.5500 b 14.49 10.95 2003.0 c 0.0366 0.0504 0.116 (7 7.905 185.9 /■ 0.0810 0.1742 excluded from consideration a type of habitat occu- pied by snow and Tanner crabs. Thus the net effi- ciency estimates we obtained may not be represen- tative of the entire area covered by the survey. Model form The form of the model that we chose to describe the relation between net efficiency and carapace width included a term (P.) that increases with width to ac- count for entry into the trawl over the footrope and another term (Pj,) that decreases with width to ac- count for entry of crabs through the belly meshes of the trawl after they have passed under the footrope. The model with both P, and P^ terms fitted the mixed- sexes data of both species significantly better than the model with only a P, term, confirming the significance of the apparent increase in net efficiency with decreas- ing size at carapace widths less than 50 mm. However, we have been unable to verify that the increased effi- ciency is indeed due to the entry of small crab through the belly meshes. Regardless of the mechanism, the important question is whether the increase in net effi- ciency at small size is a normal property of the 83-112 trawl or an artifact of the experiment. We attempted to answer this question by comparing the width distri- butions of crab captured in the trawl net of the experi- mental trawl with those captured in the standard trawl at the same or nearby stations during the survey pre- ceding the experiment, but patchiness and small sample sizes made the results equivocal. In light of this uncertainty, the composite form of the model is advantageous because it allows separa- tion of the P. and P^ effects. For example, if further research reveals that the increase in efficiency at small size is associated with the attachment of the auxiliary trawl, then an estimate of net efficiency without this effect can be obtained by deleting the P^ term and by evaluating the reduced model with the parameters from the fit of the complete model (Table 1). For this reason, the bootstrapped confi- dence intervals we provide (Fig. 4) are computed two ways, by evaluating the full model and by fitting the full model but evaluating only the P, term. However, because the P^ term decreases rapidly with carapace width, the full and reduced models predict nearly identical values of net efficiency at carapace widths above 50-60 mm, which includes the mature and commercial width ranges for both species. Efficiency differences between biological subgroups The survey trawl is less efficient for mature female Tanner crab than for the mixed-sexes group over the same range of carapace width. Since the mixed-sexes group is primarily male at these sizes, the difference in efficiency is likely due to between sex differences in both morphological features and behavior. Male Tanner crab have longer legs, in relation to their cara- pace width, than do mature females. Not only does this increase their effective size in relation to the trawl mesh but also may result in their bodies being held higher off the bottom in relation to the footrope. In addition, male Tanner crab may be caught more efficiently because they are less sedentary than fe- males and less prone to bury themselves.^ The survey trawl efficiency does not differ between mature and immature male Tanner crab. We were concerned that mature males, which have larger claws and are perhaps more active than immature males, might be preferentially selected. If such se- lection did occur, it would lead to an underestimate of the carapace width at maturity which is used in establishing minimum width limits for the commer- cial fishery. Apparently such preferential selection of mature males does not occur. In the foregoing, we estimated values of net effi- ciency or the capture probability of crab that occur between the wing-tips of the trawl net. For stock assessment purposes, however, it may be more con- venient to consider these values expressed as trawl efficiency or the capture probability of crab that oc- cur between the trawl doors. In situations where the bridles herd animals into the path of the trawl, fur- ther experiments to estimate sweep efficiency would be required to estimate trawl efficiency (Dickson, 1993a, 1993b). However, experiments on herding snow crab indicate that this species is not herded by the bridles of the 83-112 eastern trawl. ^ Thus, trawl efficiency can be estimated simply by multiplying net efficiency by the quotient of the net width divided by the door width (i.e. 0.30).2 ■* Stevens, B. 1998. Kodiak Laboratory, National Marine Fish- eries, P.O. Box 1638, Kodiak, AK. Personnal commun. 624 Fishery Bulletin 97(3), 1999 Tanner crab Mixed sexes - full model Mixed sexes - reduced model Mature females Snow crab Mixed sexes • full model 50 100 150 Mixed sexes - reduced model 150 Carapace width (mm) Figure 4 Fitted model and 95'3 bootstrap confidence intervals about the predicted value of net efficiency. For the mixed-sexes categories of both crab species, values of net efficiency are predicted in two distinct ways after fitting the full model ( Equation 1 1 to the data. First, net efficiency was predicted by evaluating both P and P^ terms of the model (top row), based on the assumption that the entry of small crabs into the trawl net through the belly mesh occurs during normal trawl operations. Second, net effi- ciency was predicted by evaluating only the P. term (i.e. setting P^=0), based on the assumption that the entry of small crab through the belly mesh was an expermental artifact caused by adding the auxiliary trawl (second row). Acknowledgments We thank J. lanelli, S. Syrjala, and D. Pengilly for reviewing the manuscript and offering helpful com- ments, P. Munro for designing the auxiliary trawl, and D. Nichol and J. Hoff for supervising the field experiment. Literature cited Alverson, D. L., and W. T. Pereyra. 1969. Demersal fish explorations in the Northeastern Pacific Ocean — an evaluation of exploratory fishing meth- ods and analytical approaches to stock size and yield forecasts. J. Fish. Res. Board Canada 26:1985- 2001. Somerton and Otto Net efficiency of a survey trawl for Chionoecetes opi/io and C bairdi 625 Armistead, C. E.. and D. G. Nichol. 1993. 1990 bottom trawl survey of the eastern Bering Sea continental shelf. U.S. Dep. Commer., NOAATech. Memo. NMFS-AFSC-7. 190p. Dickson, W. 1993a. Estimation of the capture efficiency of trawl gear. I: development of a theoretical model. Fish. Res. 16:239-253. 1993b. Estimation of the capture efficiency of trawl gear. II: testing a theoretical model. Fish. Res. 16:255-272. Efron, B., and R. Tibshirani. 1993. An introduction to the bootstrap. Chapman and Hall. New York. N^', 436p. Engas, A., and O. R. Gode. 1989. Escape of fish under the fishing line of a Norwegian sampling trawl and its influence on survey results. J. Cons. Int. Explor. Mer 45:269-276. Hilborn, R., and M. Mangel. 1997. The ecological detective: confronting models with data. Princeton Univ. Press, Princeton, NJ, 315 p. Millar. R. B. 1992. Estimating the size-selectivity of fishing gear by con- ditioning on the total catch. J. Am. Stat. Assoc. 87:962- 968. Somerton, D. A. 1980. A computer technique for estimating the size of sexual maturity in crabs. Can. J. Fish.Aquat. Sci. 37:1488-1494. Venables, W. N., and B. D. Ripley. 1994. Modern applied statistics with S-plus. Springer- Verlag, New York, NY. 462 p. Walsh, S. J. 1992. Size-dependent selection at the footgear of a ground- fish survey trawl. N. Am. J. Fish. Man. 12:625-633. 626 Abstract.— We sampled inner shelf habitat in the northeast Gulf of Mexico, for age-0 red snapper. Lutjanus cam- pechanus. to estimate growth rates and seasonality, as well as to identify nurs- ery habitats. We collected 7507 age-0 red snapper in 1994 and 1995. from 536 10-min trawl tows. Red snapper first settled to benthic habitat in June after reaching 17.4 mm standard length ( age=26 d ). In both years, catch per unit of effort (CPUE=number/10-min tow) peaked July through September, then declined in the fall as fish were leaving the habitat before winter Most fish (SO-Sl^f ) were caught at one location, 13 km south of Mobile Bay, Alabama. At this location in 1995, the August CPUE ±SE (712 ±243) far exceeded all previous estimates. Based on otolith microincrements, hatching-date fre- quencies showed distinct cohorts in June and July 1994 and May and June 1995. Growth rates for the June (0.77 mm/dt and July (0.71 mm/dl cohorts in 1994 were significantly faster com- pared with growth rates for May (0.51 mm/d ) and June (0.67 mm/d ) cohorts in 1995. Density-dependent mechanisms may be operating with faster growth rates and lower CPUEs in 1994, com- pared with slower growth rates and higher CPUEs in 1995. However, envi- ronmental constraints may also be op- erating, as indicated by the slow growth rate of the May 1995 cohort that prob- ably resulted from colder temperatures. Newly settled red snapper were aggre- gated on the inner shelf, at a particu- lar location and time period. These con- centrations indicated an important nurs- ery habitat just south of Mobile Bay, Ala- bama, from July through September. Nursery habitats, growth rates, and seasonality of age-0 red snapper, Lutjanus campechanus, in the northeast Gulf of Mexico* Stephen T. Szedlmayer Joseph Conti Marine Fish Laboratory Department of Fisheries and Allied Aquacultures Auburn University 8300 State Highway 104 Fairhope, Alabama 36532 E-mail address (for S T Szedlmayer) sszedlmaS'acesag auburn edu Manuscript accepted 20 August 1998. Fish. Bull. 97:626-635 (1999). To enhance survival some fishes spawn many times within a given season (Lambert an10%, a third blind count was made. The means of the two closest counts were then used to estimate ages. If counts still differed by >10 '^ after three counts, the otolith was rejected. Age of each fish was subtracted from date of cap- ture for hatching-date estimations. Hatching-date histograms were plotted after applying a three-day moving average for each year, and local minima were used to separate cohorts (Szedlmayer et al., 1991 ). A significance level of <0.05 was used for all statistical analysis. Catch per unit of effort (CPUE=mean num- ber/10-min tow) was estimated for each sample date and station. Prior to analysis CPUEs were log(.r-i-l) base-10 transformed. Two-way analysis of variance (AN OVA) was used to test for significant differences in mean CPUEs and mean SLs among dates and sta- tions. Waller-Duncan's multiple comparison test was used to show specific differences detected by the ANOVAs, Growth rates were estimated by linear regression of SL on collection date, and SL on age from otolith microincrement counts. Analysis of covariance (ANCOVAl, Student's t-test, and Tukey's multiple comparison test were used to compare estimated growth rates (Zar, 1984). Results Salinity ranged from 30.5 to 35.4 ppt, dissolved oxy- gen from 3.7 to 7.9 ppm in 1994. Temperature in- creased from 22°C in June to a peak of 29°C in Au- gust, then decreased to 17°C by January 1995. The temperature (26°C) in early June 1995 was higher compared with the previous year, stayed near this level through July, increased to 30°C in September, then decreased in the fall (Fig. 2). Temperature data from the NOAA buoy indicated that temperatures 628 Fishery Bulletin 97(3), 1999 were probably colder in May of both years at the sample sites (Fig. 2). Temperature, salinity, and dis- solved oxygen were similar among stations (Table 1). Visual observations of debris in trawl samples indicated shell deposits at station 1 that were not apparent at other stations. Grain size analysis indi- cated similar sediments between stations 1 and 5, and between stations 2 and 4, whereas station 3 had a significantly higher silt-clay fraction compared with all other stations (ANOVA, P<0.05; Table 1). From these nursery habitats, we collected 7507 age-0 red snapper from 536 10-min trawl tows in 1994 and 1995. Significantly more fish were collected in 1995 than in 1994 (ANOVA, P<0.05; Fig. 3). Age-0 red snapper first settled to these habitats in late June, showed highest abundance July through Sep- tember, then steadily declined in the fall of both years (ANOVA, P<0.05; F'ig. 3). Most fish, 80-81%, were collected at station 1 (ANOVA, P<0.05: Fig. 4). At station 1, significantly higher peaks in CPUE ±SE for 1994, were 63.6 ±8.9 in July, 60.8 ±26.2 in Au- gust, and 40.3 ±13.4 in September (ANOVA, P<0.05). In late August 1995 at station 1 , we observed a CPUE (712 ±243) that far exceeded all previous estimates from this study or any previous study. Other signifi- cant peaks in CPUE for 1995 at station 1 were 76.2 ±21.4 in mid-August, and 81.4 5.0 in mid-September (ANOVA, P<0.05). Age-0 red snapper ranged from 17.8 to 124.4 mm SL. Fish first settled from the plankton after they reached 17.8 mm SL. The smallest (<20 mm SL) were present both years into mid-September, after which no new settlers were detected (Fig. 5). Significantly higher mean SLs were detected by late August in both years compared with earlier sample periods; size significantly increased with season (ANOVA, P<0.05; Table 2; Fig. 5). Fish were significantly larger earlier in 1995, but fish from 1994 caught up in size by Sep- tember, afl:er which 1994 fish were significantly larger than 1995 fish (ANOVA, P<0.05; Table 2; Fig. 5). In 1994, age-0 red snapper were first abundant (CPUE=63.6) at station 1 and had limited settlement (CPUE=2.0) at station 3 (Fig. 6). Also, fish were sig- nificantly larger (ANOVA, P<0.05), less abundant, and showed up later at all other stations compared with fish at station 1 in 1994. We did not detect pat- terns in inshore-offshore movement from fish size, location, and seasonality; we did determine that af- ter first settlement at station 1, fish showed expan- sion in all directions (Fig. 6). In 1995, new recruits were most abundant in July at station 3 (CPUE=78). After July, patterns were similar to 1994, and most fish were observed at station 1 and fewer larger fish were observed at other stations later in the season (Fig. 7). \ ] \ 1 1 i 1 1 1 MJJASONDJ Month Figure 2 Bottom water temperature ( C). salinity (ppt), and dis- solved oxygen (ppml pooled by stations, in the northeast Gulf of Mexico, from June 1994 to January 199.5, and June 1995 to January 1996. Lines without symbols are mean daily surface temperatures based on 10-d moving averages, obtained from a NOAA moored buoy. The required precision (<10% ), was shown for 57% of all otolith counts (Table 3). Age of red snapper was 26 to 144 days; thus fish may spend up to four months in these habitats. We detected separate late May-June and July cohorts in 1994, and May and June cohorts in 1995, ft-om hatching-date frequencies (Fig. 8). Growth rates were significantly different among cohorts ( ANCOVA, P<0.05). The fastest growth rates were observed for June (0.77 mm/d) and July (0.71 mm/d( cohorts in 1994. Growth rates for May (0.51 mm/d) and June (0.67 mm/d) cohorts in 1995 were significantly less than the previous year and signifi- cantly different from each other (ANCOVA, P<0.05; Fig. 9). Growth rate estimates from SL on varying dates (0.52 mm/d in 1994; 0.62 mm/d in 1995; Fig. 5) were significantly lower than most cohort growth estimates, with the exception of the slow growing May 1995 cohort (ANCOVA, P<0.05). In station com- parisons for 1995, fish from station 2 had a signifi- cantly faster growth rate (0.86 mm/d), followed by fish from station 1 (0.71 mm/d), whereas slower growth rates were observed for station 3 (0.54 mm/ d) and station 5 (0.60 mm/d; ANCOVA, P<0.05; Table Szedlmayer and Conti: Nursery habitat, growths rates, and seasonality of age-0 Lut/anus campechanus 629 Table 1 Mean temperat northeast Gulf ures, mean salinities, of Mexico. See Figure mean dissolved oxygen 1, for station locations. IDO), depths and +SEs substr ate types for trawl stations from the Station Mean temperature (°C) Mean salinity (ppt) Mean DO (ppm) Depth (m) Substrate type 1994 1 24.5+1.0 32.7+0.5 5.4+0.4 18.7+0.6 fine sand-shell 9 24.6+1.3 29.9+0.9 6.1+0.5 15.6+0.6 coarse sand 3 24.3+1.4 31.7+0.7 6.1+0.5 13.4+0.6 silt, clay, sand 4 24.9+1.3 32.3+0.5 6.3+0.4 15.0+0.4 coarse sand 5 24.6+1.2 32.7+0.5 6.3+0.4 20.7+0.6 fine sand 1995 1 24.0+1.2 35.5+0.7 6.3+0.4 19.5+0.1 fine sand-shell 2 25.3+1.2 32.6+0.4 6.2+0.2 16.1+0.7 coarse sand 3 22.6+1.3 33.6+1.3 5.5+0.5 12.7+0.2 silt, clay, sand 4 23.6+1.2 35.6+0.4 6.1+0.2 14.8+0.2 coarse sand 5 23.8+1.3 34.8+0.9 6.5+0.3 20.8+0.2 fine sand 4). Station 4 growth rates were not estimated be- cause of low catch rates. Discussion Age-0 red snapper used inner shelf habitat at first settlement, and an area 13 km south of Mobile Bay (station 1) held particularly high numbers. No other study has reported such high CPUEs of age-0 red snapper. Many factors could contribute to this important nursery habitat at station 1. Temperature, dissolved oxygen, salinity, and depths, showed little difference among stations, but one potential aspect that may provide avenues for future studies is substrate type. The substrate in this area may be highly suitable, e.g. relic shell beds (Schroeder et al., 1995) with fine sand sedi- ments, and there is a general shift from shell-sand- mud to "clean" sand east of Mobile Bay.' where few fish were collected. Age-0 red snapper may be selecting such shell habitats. For example, age-0 red snapper showed a preference for shell over sand substrata in laboratory studies (Szedlmayer and Howe, 1997 land a preference for almost any small relief structure (e.g. shells) in relation to flat sand substrata in natural habitats (Workman and Foster, 1994). Shells were observed in trawl col- lections at station 1 and not apparent in trawl samples at other stations. However, we were un- ' Hummell, R. L. 1998. Geological survey of Alabama, Tuscaloosa. AL. Personal commun. a. o 25 20 15 10 - 5 - 1994 CPUE all stations 160 n 140 120 100 80 60 40 20 0 -1* •r- 1 ] ; -I - I* r Jun Jul Aug Sep Oct Nov Dec Jan Month Figure 3 Mean CPUE lnumber/10-min towi of age-0 red snapper, Liitjaniif: compechaniis. by date from the northeast Gulf of Mexico. Each point represents mean CPUE for that date. Er- ror bars are standard errors. Different letters show signifi- cant differences both within and between years (ANOVA; Waller-Duncan test; P<0.05). 630 Fishery Bulletin 97(3), 1999 able to confirm the presence of shell deposits from sediments collected by SCUBA divers. Red snapper first settled in June, showed peaks in July, August, and September, then quickly declined Table 2 MeanSL(nim)ofage-Ored snapper Lutjanus campechanus. for each sample date in 1994 and 1995, from the northeast | Gulf of Mexico. Means with different letters sho w signifi- cant differences (ANOVA, Waller- Duncan's test; P<0.05). Mean SL Date n (mm) SE 30 June 94 5 25.0" 2.01 11 July 94 328 24.1" 0.19 1 August 94 121 36.1' 0.64 U August 94 137 26.6" 0.72 17 August 94 401 36.1' 0.66 30 August 94 165 41.6'' 1.22 8 September 94 322 42.9'' 0.46 26 September 94 154 70.5''B 1.88 13 October 94 72 75.4' 1.59 31 October 94 54 85.91 1.28 14 December 94 42 99.2'' 3.12 10 January 94 2 116.1' 8.35 11 July 95 108 33.2'"' 0.62 24 July 95 454 31.8'' 0.40 9 August 95 485 35.5'" 0.35 25 August 95 3696 41. l"" 0.08 12 September 95 572 57. r 0.38 22 September 95 259 66.5''« 0.66 25 October 95 53 60.8'- 2.36 14 November 95 44 72. yh. 2.52 2 December 95 18 65.3f 4.83 10 January 96 15 89.4) 4.67 Table 3 Comparison o estimation of campechanus. "otoliths prepared and percent accepted for age in age-0 red snapper, Lutjanua Date Otoliths prepared Otoliths with >10% precision Percent accepted 11 Jul 94 318 172 54 17 Aug 94 316 149 47 8 Sep 94 322 118 37 10-13 Jul 95 108 78 72 9-10 Aug 95 245 I'.SO 73 11-13 Sep 95 279 160 57 Total 158S 857 57 in numbers from the habitat. There is evidence that this decline of age-0 red snapper from open relatively fiat habitats in the fall results from larger age-0 fish migi'ating to reef structure. In a study offish recruit- ment to 1-m'^ concrete artificial reefs, 20 km south of Mobile Bay, age-0 red snapper showed large (up to 89 fish/reef) influxes to these reefs in September and October.- This seasonal pattern was similar to the pattern suggested by Holt and Arnold ( 1982). They found that recruitment of small red snapper to benthic substrate occurred primarily from June through September, and in the fall a few larger age-0 fish were collected by trap near a sunken ship.'* Metamorphosis from larva to juvenile is ecologi- cally important in the early life history of fish. In - Szedlmayor, S. T. 1994. Production or attraction: an evalu- ation of artificial reefs in coastal Alabama. Alabama Univer- sities Tennessee Valley Research Consortium Final Report, Au- burn University, Auburn, AL, 48 p. 3 Holt, S. A. 1997. Marine Science Institute, Univ. Texas, Port Aransas, TX. Personal commun. Figure 4 Mean CPUE lnumber/10-min tow) of age-0 red snapper, Lutjanus campechanus, by station from the northeast Gulf of Mexico. Error bars are standard errors, and n's are num- ber of trawl tows. Different letters show significant differ- ences both within and between years (ANOVA; Waller- Duncan test; P<0.05). Szedlmayer and Conti: Nursery habitat, growths rates, and seasonality of age-0 Lut/anus campechanus 631 120 n 100 80 60 40 i ^° C •o s •D C o5 120 100 H 80 60 40 20 1994 n = 1803 /•-square = 0 62 b = 0.52 mm/d 1995 n = 5704 r-square = 0 52 b = 0.43 mm perd I I I 1 1 1 1 1 Jun Jul Aug Sep Oct Nov Dec Jan t^onth Figure 5 Linear regression of age-0 red snapper, Lutjanus campechanus. SL on date for 1994 and 1995, for all fish collected from the north- east Gulf of Mexico. Data points are individual fish. Growth rates were significantly different between years (Mest; P<0.05). many fish species this transition occurs simulta- neously with habitat change, i.e, from pelagic larvae to benthic juveniles. On the basis of smallest fish collected during this study, larval red snapper meta- morphose near 26 d and at 18 mm in length. Collins et al. (1980) suggested similar settlement sizes (12.4 to 22.4 mm SL) from plankton net samples. Age-0 red snapper otoliths had daily microin- crements when growth rates were >0.3 mm/d (Szedl- mayer, 1998). In the present study, the nursery pe- riod was in the warmer summer months, when food should be abundant; and by all indications, growth rate far exceeded the 0.3 mm/d threshold rate. Thus, back calculation of first increment formation accu- rately estimated hatching date for age-0 red snap- per. One difficulty with the otolith ageing methods was the low (579'^) percentage of otoliths that met the required predetermined precision. However, this low percentage was due to personnel skill in otolith Table 4 Comparison of growth rates among selected 1995. Means with different letters show signif ences (ANCOVA, Tukey's test; P<0.05). stations in cant differ- Station Growth rate mm/d n r2 1 0.7P 227 0.86 2 0.86'' 25 0.89 .3 0.54' 115 0.80 5 oec^ 49 0.76 preparation. For example, an inexperienced techni- cian would typically over polish several otoliths, that were subsequently rejected when counting. This same technician, after gaining experience and in- 632 Fishery Bulletin 97(3), 1999 4U- IIJul 20- A A 318 AB 10 pK- 1 1 1 1 1 1 20- D 83 BC 32 1 Aug 1 1 1 1 1 1 20- AB 137 11 Au8 Not sampled u - 1 1 I 1 1 1 40- T 20- CD 316 E 74 17 Aug n 1 1 1 1 1 1 60- 30 Aug T r^ 40- FG 10 FG 26 20- D 128 n 1 1 1 1 1 1 40- 20- E 322 esep Not sampled n 1 1 1 1 1 1 100 80 60H 40 20-1 0 80- 60- 40 20- 0- 80 60-1 40 20 0 100 80 60 40 20 0 Stations (1994) Figure 6 Mean standard lengths by station and date in 1994 for all collections that had >10 fish. Numbers in bars are number offish measured. Different letters show significant differences among stations and dates ( ANOVA; Waller-Duncan test; P<0.05). 26 Sep creased skill, would make an acceptable preparation of the second otolith from the same fish. Hatching-date distributions clearly indicated sepa- rate hatching cohorts with peaks approximately one month apart. The multiple spawnings within a given year in red snapper may be an adaptation to increase survival. By spreading reproductive efforts over time, a species may increase the probability of matching the correct biotic and abiotic conditions for enhanced survival and subsequent recruitment. Two differences were apparent when hatching-date estimates were compared with estimated spawning periods from gonadosomatic index (GSI) maxima (Collins et al., 1996). First, GSI estimates showed little indication of May spawning, as suggested by May hatching dates in the present study; and sec- ond, present estimates showed little indication of August spawning, as indicated by high GSIs for red snapper in August (Collins et al. 1996). These differ- ences between hatch dates and GSIs were probably Szedlmayer and Conti: Nursery habitat, growths rates, and seasonality of age-0 Lutianus campechanus 633 40 n 20 0 20 0 40 20 0 60 40 20 0 60 40- 20- 0 80 H 60 40 H 20 0 "1 ^r 11 Jul \ 1 80 24 Jul 60 9 Aug 1 1 1 40 20 H 0 1 80 25 Aug FG 3561 12 Sep 22 Sep MN I 68 60- 40 20 H -1 0 60 40 20 r^— 1 0 80 - 60 40 20 "T r T r "I I r 12345 12345 Stations (1995) Figure 7 Mean standard lengths by station and date in 1995 for all collections that had >10 fish. Numbers in bars are number offish measured. Different letters show significant differences among stations and dates ( ANOVA; Waller-Duncan test; P<0.05). due to yearly variation. Collins et al. ( 1996) sampled from 1991 to 1993. Lack of evidence for August spawn- ing in our study may also have resulted from low survival of later spawned fish or fi"om new recruits hav- ing settled to different habitats not sampled in the present study The June and July 1994 cohorts had growth rates that were significantly faster in relation to 1995 co- hort growth rates. Growth rates estimated from SL on varying dates were probably inaccurate owing to extended settlement of new recruits, but results showed the same pattern as SL on age estimates, i.e. 1994 growth rates were significantly faster com- pared with 1995 estimates (^test, P<0.05; Fig. 5). If growth rates were an indication of the probability of survival (Houde, 1989; Sogard, 1997), then 1994 co- horts may have had an advantage over the 1995 spawned cohorts. Growth rate typically decreases with decreasing temperature and may have resulted in the low growth rate observed for the May 1995 cohort. Thus, it appears that both density dependent and density independent mechanisms are operating 634 Fishery Bulletin 97(3), 1999 20 15 - 10 -J June cohort July cotiort n = 439 c 0 20 June cohort May June Hatching date July Figure 8 Hatch date frequencies of age-0 red snapper, Lutjanus campechanus. pooled over all dates from 1994 and 1995, A three day moving average was applied before plotting. (Bromley, 1989; Rijnsdorp and Vanleeuwen, 1992; Rogers, 1994). In 1994, there were fewer fish but faster growth rates than those in 1995, and in 1995 the early May cohort showed the slowest growth rates, probably from colder temperatures. This scenario suggests that if red snapper spawn early (May), temperature may play an important role. As the season progresses into July and August, temperature increases, settlement increases, and competition for food and shelter may become the limiting factors, such that later spawnings in August have lower survival. The inner shelf habitat in the northeast Gulf of Mexico, particularly the location 13 km south of Mobile Bay, Alabama, should be considered an im- portant nursery area for age-0 red snapper from July through September. Future research should address mechanisms that help to define this red snap- per nursery habitat, and back calculation methods could be used to compare the relative contributions of various within-year cohorts to subsequent year classes. y^ 80 - ' y^ yy yiy 60 - O SLTc'^t-^* • ' OrctPgSi^' 40 - OoMjSJwS ^^^Seos • June 1994 cohort Ri&ffi»o° n = 207. b = 0 77 r-square = 0 95 jJBP^'"" a _ 20 - '^^f^ ' 0 July 1994 cohort E g //^ n = 232 b = 0 71 r-square = 0 81 ^ CT a 1 0 - I 1 1 p ■D o 1 80 - c to 55 Sa oo ^^ .••^^« 60 ~ 0 °o 3'f°^J^,itf^ ° 0<%??\<^^r?. n 40 - °o^^^' '° °Wf^''- • May 1995 cohort o ,^ares?'°5 • n = 98 b = 0 51 Q^^Mpfi ° r-square = 0 85 20 - C o June 1995 cohort ^ n = 318 b = 068 r-square = 0 83 0 - b III 20 40 50 80 100 120 Age (d) Figure 9 Growth rates of separate cohorts of age-0 red snapper. Lutjaniia campechanus. from SL on age regressions from the northeast Gulf of Mexico. Data points are individual fish. Different letters show significant differences among cohorts both within and be- tween years (ANCOVA; Tukey's test; P<0.05). Acknowledgments We thank Jason Lee for help in field collections and otolith preparation. We also thank Karen Belcolore for data entry and for typing the manuscript. This study was funded by NOAA, NMFS, MARFIN grant number USDC-NA47FF0018-0. Literature cited Baughman, J. L. 1943. The lutjanid fishes of Texas. Copeia 43:212-15. Bradley, E., and C. E. Bryan. 1975. Life history and fishery of the red snapper I Lu//a«u.<: campechanus) in the northwestern Gulf of Mexico: 1970- 1974. Proc. 27th Annu. Gulf Caribb. Fish. Inst, and 17th Annu. Int. Game Fish Res. Conf 27:77-106. Szedlmayer and Conti Nursery habitat, growths rates, and seasonality of age-0 Lutjonus campechanus 635 Bromley, P. J. 1989. Evidence for density-dependent growth in North Sea gadoids. J. Fish Biol. 35:117-123, Collins, L. A., J. H. Finucane, and L. E. Barger. 1980. Description of larval and juvenile red snapper. Lutjanus campechanus. Fish. Bull. 77:965-974. Collins, L. A., A. G. Johnson, and C. P. Keim. 1996. Spawning and annual fecundity of the red snapper {Lutjanus campechanus) from the northeastern Gulf of Mexico. In F. Arreguin-Sanchez, J. L. Munro, M. C. Balgos, and D. Pauly (eds. ), Biology, fisheries and culture of tropical groupers and snappers, p. 174-188. ICLARM Conf Proc. 48. Makati City. Philippines. Holme, N. A., and A. D. Mclntyre. 1971. Methods for the study of marine benthos. IBP hand- book no. 16. Blackwell Scientific Publications, Oxford, 334 p. Holt, S. A., and C. R. Arnold. 1982. Growth ofjuvenile red snapper Lu(/ari(/srampcc/ia;i us, in the northwestern Gulf of Mexico. Fish. Bull. 80:644-648. Houde E. D. 1987. Fish early life dynamics and recruitment variability. Am. Fish. Soc. Sym. 2:17-29. 1989. Subtleties and episodes in the early life of fishes. J. Fish Biol. 35:29-38. Lambert. T. C, and D. M. Ware. 1984. Reproductive strategies of demersal and pelagic spawning fish. Can. J. Fish. Aquat. Sci. 41:1565-1569. Miranda, L. E., and W. D. Hubbard. 1994. Length-dependent winter survival and lipid compo- sition of age-0 largemouth bass in Bay Springs Reservoir, Mississippi. Trans. Am. Fish. Soc. 123:80-87. Moseley, F. N. 1966. Biology of the red snapper, Lutjanus aya Bloch, of the northwestern Gulf of Mexico. Publ. Inst. Mar Sci. Univ. Texas 11:90-101. Ogren, L. H., and H. A. Brusher. 1977. The distribution and abundance of fishes caught with a trawl in the St. Andrew Bay system, Florida, North- east Gulf Science 1)2):83-105. Rijnsdorp, A. D., and P. I. Vanleeuwen. 1992. Density-dependent and independent changes in so- matic growth of female North Sea plaice, Pleuronectes platessa. between 1930 and 1985 as revealed by back-cal- culation of otoliths. Mar. Ecol. Prog. Sen 88:19-32. Rogers, S. I. 1994. Population density and growth rate of juvenile sole Solea solea (L). Neth. J. Sea Res. 32:353-360. Schroeder, W. W., A. W. Shultz, and O. H. Pilkey. 1995. Late quaternary oyster shells and sea-level history, inner shelf, northeast Gulf of Mexico. J. Coast. Res. 11:664-674. Secor, D. H., J. M. Dean, and E. H. Laban. 1991. Manual for otolith removal and preparation for mi- crostructural examination. Electric Power Res. Inst, and Belle W. Baruch Inst. Mar. Biol. Coast, Res., Columbia, SC, 85 p. Sogard, S. M. 1997. Size-selective mortality in the juvenile stage of te- leost fishes: a review. Bull. Mar Sci. 60:1129-1157. Spurr, A. R. 1969. A low-viscosity epoxy resin embedding medium for electron microscopy. J. Ultrastruct. Res. 26:31-43. Szedlmayer, S. T. 1998. Comparison of growth rate and formation of otolith increments in age-0 red snapper J. Fish Biol. 52:58-65. Szedlmayer, S. T., and J. C. Howe. 1997. Substrate preference in age-0 red snapper. Lutjanus campechanus. Environ. Biol. Fish. 50:203-207. Szedlmayer, S. T., and R. L. Shipp. 1994. Movement and growth of red snapper, Lutjanus campechanus, from an artificial reef area in the northeast- ern Gulf of Mexico. Bull. Mar Sci. 55:887-896. Szedlmayer, S. T., M. M. Szedlmayer, and M. E. Sieracki. 1991. Automated enumeration by computer digitization of age-0 weakfish Cynoscwn regalis scale circuli. Fish. Bull. 89:337-340. Szedlmayer, S. T., M. P. Weinstein, and J. A. Musick. 1990. Differential growth among cohorts of age-0 weakfish Cynoscwn regalis in Chesapeake Bay. Fish. Bull. 88:745- 752. Workman, I. K., and D. G. Foster. 1994. Occurrence and behavior of juvenile red snapper, Lutjanus campechanus, on commercial shrimp fishing grounds in the northeastern Gulf of Mexico. Mar Fish. Rev 56:9-11. Zar, J. H. 1984. Biostatistical analysis, 2nd edition. Prentice Hall, Inc., Englewood Cliffs, NJ, 718 p. 636 Abstract.— This study examined the relation between statolith and somatic growth in the tropical squid Sepioteuthis lessoniana. Five separate linear dimen- sions were measured on the statoliths of 103 individuals ( 17-245 mm mantle length 1. In addition the statoliths of 80 adults (82-245 mm mantle length) were weighed. Statolith increment analysis provided age estimates for 78 individuals. Statolith total length was cprrelated with age for squid less than -60 days of age, although neither sta- tolith total length nor weight was a useful predictor of age in older squid. Combining the five statolith dimen- sions to produce a description of sta- tolith shape provided only slightly bet- ter estimates of age than statolith to- tal length or weight alone. Statolith shape changed during ontogeny, devel- oping from relatively elongate juvenile statoliths into the adult form with more robust dorsal and lateral domes. This development was reflected in wider spacing and superior optical definition of daily growth increments in the dor- sal and lateral domes of adult sta- toliths, in relation to the slower grow- ing rostrum. Growth of S. lessoniana statoliths does not appear to be strongly linked to mantle growth; both statolith total length and weight increase more slowly than mantle length. Ontogenetic changes in size and shape of statoliths: implications for age and growth of the short-lived tropical squid Sepioteuthis lessoniana (Cephalopoda: Loliginidae) Ross Thomas Natalie A. Moltschaniwskyj School of Marine Biology and Aquaculture James Cook University Townsville, Queensland 4811, Australia Present address (for N A Moltschaniwskyi, contact author): School of Aquaculture University of Tasmania PO Box 1214 Launceston, Tasmania 7250. Australia. E-mail (for N A Moitscfianiwskyj, contact author) natalie moltschaniwskyi aulas edu au Manuscript accepted 12 August 1998. Fish. Bull. 97:636-645 (1999). Knowledge of the age structure and growth rates in naturally occurring populations is fundamental to esti- mating demographic parameters and evaluating ecological processes. Periodic growth increments in squid statoliths are a reliable and accu- rate tool for determining the age structure of squid populations (Villanueva, 1992;Ai-khipkin, 1993; Bigelow, 1994; Jackson, 1994). Growth-rate calculations based on statolith age estimates indicate that tropical squid are short-lived and gi-ow continuously throughout their lifespan (Jackson, 1990; Jackson and Choat, 1992). Growth rates of tropical squid can be strongly influ- enced by changes in their environ- ment, including temperature fluc- tuations (Forsythe and Hanlon, 1989), and availability of food (Forsythe, 1993). Consequently, growth rates, final size, and re- sponse to changing conditions of tropical squid may vary greatly within or between species (Jackson, 1990; Jackson and Choat, 1992). Apart from obtaining size-at-age information from statoliths, there is also the potential to obtain ecologi- cal information on past growth his- tories of some squid species (Jack- son, 1994). Any attempts made at constructing growth histories of squid based on statolith microstruc- ture will benefit greatly from a thor- ough understanding of growth and development of the statolith and how this relates to growth of so- matic tissue. Daily growth increments in fish otoliths, which are analogous to those in squid statoliths (Radtke, 1983 ), provide valuable data on the age structure and growth rates of exploited fish species (Campana and Neilson, 1985; Jones, 1986; Stevenson and Campana, 1992). Recent studies of fish somatic- otolith growth relationships may contribute to investigations of so- matic-statolith growth relation- ships in squid. Evidence exists sup- porting the direct relationship be- tween age and otolith weight for several fish species (Boehlert, 1985; Fletcher, 1991), although further validation is required before otolith weight can be used reliably to esti- mate age. Several authors have also demonstrated that slower-growing fish have larger otoliths than simi- lar sized, faster-growing individu- als (Mosegaard et al., 1988; Secor and Dean, 1989; Wright et al., 1990; Thomas and Moltschaniwskyi Ontogenetic changes in size and shape of statoliths of Sepioteuthis lessoniana 637 Mugiya and Tanaka, 1992 ), indicating a possible dis- association between otolith and somatic growth. Similarly, disassociation between statolith and so- matic tissue growth has been demonstrated in the tropical loliginid squid Loligo chiiieiisis (Jackson, 1995), and also in two ommastrephid squids, Todarodes angolensis and Todaropsis eblanae (Lipinski etal., 1993). The periodic growth increments in squid statoliths are bipartite structures consisting of a discontinu- ous zone and an incremental zone comprising arago- nite (CaCO.^) crystals (Lipinski, 1986). Specific mor- phological differences in the aragonite crystal struc- ture have been illustrated between statolith regions within several squid species (Lipinski, 1993). Crys- tals of the lateral dome region often display greater variation in shape, size, and orientation than the more homogenous wing crystals (Lipinski, 1993). These variations in crystal structure suggest that accretion is not uniform and that different regions of the statolith may grow at different rates. The present study provides an assessment of the growth and shape of Sepioteuthis lessoniana sta- toliths based on the linear measurement of statolith dimensions. The potential for using statolith shape descriptions as a proxy for age estimations and the relation between statolith and somatic tissue growth in these squid are examined. Overall descriptions of individual statolith shape covered a broad length and weight range of squid, providing a basis for compari- son with age estimates determined by using tradi- tional analysis techniques. With the increased use of statoliths in squid age and growth studies, it is important to understand how statoliths grow and how this growth relates to growth of somatic tissue. Materials and methods Study species and collection methods A total of 103 Sepioteuthis lessoniana individuals were captured from waters around the Townsville region of north Queensland, Australia, between Janu- ary and August 1995. Twenty-three juvenile squid (16-43 mm mantle length [ML]) were captured by using purse seines and dip nets. Juveniles often shel- ter in surface waters among floating debris or artifi- cially constructed shade devices and are easily net- ted. Eighty adult squid (82-245 mm ML) were cap- tured from coastal waters at night with squid jigs and by means of light attraction. Individuals were separated into two age groups for these descriptions: juvenile (<60 days) and adult (>60 days). This sepa- ration was based on the techniques used to capture Figure 1 Sepioteuthis lessoniana statolith showing the dimensions measured for morphometric analy- ses. TL = total statolith length; DLL = dorsolat- eral length; VLL = ventrolateral length; LDL = lateral dome length; MW = maximum width. the individuals, which may be related to differences in life style and ecology. Size measurements Mantle length was measured from all Sepioteuthis lessoniana specimens upon capture. Wet weight was not obtained for the majority of individuals because of the inability to weigh animals accurately on a boat. Statoliths were removed and stored in 707^ alcohol, and adult statoliths were later weighed to the near- est 0.01 mg. Weights of juvenile statoliths were not recorded because they were too small to obtain accu- rate measurements. Five dimensions were measured from each statolith following the descriptions of sta- tolith shape by Clarke (1978): total length, dorsolat- eral length, ventrolateral length, lateral dome length, and maximum width (Fig. 1 ). Dimensions were mea- sured from whole statoliths viewed with the ante- rior (concave) side positioned upward by using an Ikegami-290 high-resolution black and white video camera mounted on a compound microscope. Squid ages were estimated by using statolith increment analysis on ground and polished statoliths (Jackson, 1990). All increment counts were made in the dorsal-dome re- gion of statoliths because increment definition was con- sistently clearest in this area. Age estimates for 62 638 Fishery Bulletin 97(3), 1999 adult and 16 juvenile squid were calculated from a mean of three separate counts made by one reader. Statistical analyses Paired-sample f-tests indicated no significant differ- ence between left and right statolith total length mea- sures (^=0.67, df=101, P=0.50, n^l02), or adult statolith weight measures it^OAT, df=79, P=0.64, n=80), but wherever possible, the left statolith was used for analy- ses. For the purpose of this study, statolith "size" refers to the length of each statolith dimension, whereas "shape" indicates that all statolith dimensions have been combined to describe overall statolith structure. Rates of growth calculated from size-at-age data were not significantly different between adult males and females (Table 1). A size comparison of males and females, adjusted for age, found no significant difference in size between the sexes (Table 2). Indi- viduals were therefore classified as either juveniles or adults, with no distinction made between adult sexes. This classification avoided over-complicating analyses and also provided a larger adult sample size. Principal components analysis (PCA) was per- formed on statolith dimensions with the covariance matrix of transformed (logj^) data, allowing complete descriptions of statolith shape for all dimensions si- multaneously. The magnitude of coefficients in the Linear regression and female Sepio of the slope. ML= Table 1 equations for the size-at-age data of male teiilhis lessoniana. SE^ = standard error mantle length (mm I. H^ for t-test: 6 = 0. Sex n Equation SE, r^ t P Male Female 32 31 age age = 0.08ML -f 5.02 = 0.08ML + 5.44 0.02 0.02 0.44 4.84 0.43 4,64 0.0001 0.0001 first eigenvector varies because of changing statolith proportions and relative growth patterns. These co- efficients indicate the nature of allometric growth for each variable (dimension) (Jolicoeur, 1963). The coefficients for each variable in the first eigenvector describe the relative growth rates of all the compo- nents simultaneously (Shea, 1985). Variables with coefficient scores on the first principal component vector equal to the mean coefficient (( l/p)" ■^, where p is the number of variables in the analysis) are iso- metric. Positive and negative allometry is indicated by values greater than and less than the mean coef- ficient, respectively (Jolicoeur, 1963). The resampling technique jack-knifing provided means and standard errors (SE) of each coefficient (Marcus. 1990). The probability that the mean coef- ficient was significantly different from the hypoth- esized value (Marcus, 1990) is \mean coefficient - hypothesized coefficient^ > T X SE X mean coefficient (1) where T = the number of standard errors for which the probability statement is made. In this case T = AAl provides a probability of 0.05. Simply, the difference between the mean coefficient and the hypothesized coefficient is greater than a critical difference (T x SE x calculated coefficient) with a probability <\IT ^. Results Statolith size relationships The smallest statolith total length (648 nm) was ob- served in the youngest specimen ( 17 days old). The largest adult statolith total length (2167 ^m) was observed in a 162-day-old mature female, although Analysis of covariance table comparing mantle 1 sngth Table 2 (ML) of male and fema e Scpioteuthis lessoniana using age as a covariate. Source df Sum of squares Mean square F-value P>F Test for equal slopes Sex Age (covariate) Sex X age (equal slopes) 1 1 1 1.46 329.60 1.20 1.46 329.60 1.20 0.32 71.21 0.26 0.5764 0.0001 0.6128 Test for differences in ML Sex Age (covariate) 1 1 0.40 333.23 0.40 333.23 0.09 73.01 0.7681 0.0001 Thomas and Moltschaniwskyi: Ontogenetic changes in size and shape of statoliths of Sepioteuthis lessoniana 639 the oldest individual was a 186-day-old mature fe- male. Adult statolith weights ranged from 0.29 mg in a 72-day-old immature male to 1.74 mg in a 144- day-old mature female. Adult statolith weight was not a reliable predictor of squid age, with only 579c (n=62) of the variation in adult ages attributable to weight of the statoliths (Fig. 2A). The strength of the relation between age and statolith total length differed between juvenile and adult squid (Fig. 2B). For juvenile squid, 859c (n = 16) of the variation in age was attributable to dif- ferences in statolith total length, indicating statolith total length maintains a close correlation with age during the first 60 days of squid growth. In adult squid, the ability to predict age based on statolith total length is poor, and only 579c (n =62) of the varia- tion in squid ages can be explained by statolith length. Statolith total length grew relatively slower than ML for both adult and juvenile squid (Fig. 3A). Adult statolith weight displayed positive allometric growth with ML (Fig. 3B). 200 160 -I 120 80 40 0 age = 63.42x + 62.70 r-=0.57, n = b2 0.0 0,4 0.8 1.2 1.6 2.0 Statolith weight (mg) 200 160 Jage-24.27e0-«'"^ .,^. r2 = 0.57, n = 62 " "^ B 20^ •i 5^- 80 - « .^ 40-. J^ge = 2.10 0.003X 0 A — r2 = 0.85, n = 16 0 600 1,200 1,800 2,400 Statolith total length lum) Figure 2 Relationship between (A) adult statolith weight and age, and iB) statolith total length and age for Sepioteuthis lessoniana specimens. Circles = adults; squares = juveniles. Changes in statolith shape The first PCA axis described 98.6% of the variation among individuals, which is due to statolith shape or the size of all statolith dimensions combined. If we use the PCA score on the first axis as a descriptor of statolith shape, we can account for only 99c and 7% of the variation in statolith total length of juve- nile (n = 16) and adult (??=78) squid, respectively (Fig. 4B). This result confirms that the first PCA axis con- tains statolith shape information and not predomi- nantly size information. Growth of the statolith along the total length, ventrolateral and lateral dome axes and in total length was relatively slow, as evidenced by negative allometric growth (Table 3). In contrast, gi-owth of the statolith along the dorsolateral axis and in total width grew relatively faster, as evidenced by the positive allometric growth (Table 3). This find- ing indicates that the statoliths of juvenile squid are relatively thin and elongate, becoming comparatively 3.4 3.2 y=2.74 + 0.43x (95%CL = 0.39-0. r2 = 0. 61, n = SO s 3.0 3.0 -I 2.8 -^"J /■ y = 2.71 + 0.45x r2 = 0.92, n = 2S !.CL = 0.41-0.49) 1.5 2.0 2.5 0.4 0.2 0.0 -0.2 ■0.4 ■0.6 y = 0.17.i-1.49x (95%CL = 1.34-1.67) !^ = 0.52, n = 80 1.0 1.5 2.0 2.5 Log matitle length (mm) Figure 3 Relation between (A) mantle length and statolith total length, and (B) mantle length and adult statolith weight for Sepioteuthis lessoniana specimens. Circles = adults; squares = juveniles. CL = confi- dence limits of the slope. 640 Fishery Bulletin 97(3), 1999 2,400 n , 2,000 - E f 1,600 - 1 1,200 - k' 2 800 -■^ 2 400 H 2 Cn 0 1 1 I 1 -1.0 -0.5 0.0 0.5 1.0 200 160 log(agc)=l,6Sx , B -I- 4.49 ^ S' 120 - 1 80 - ■ 40 - ••^rogtage) = 2.42x + 4.49 r2=0.73,«=y5 -1 .0 -0.5 0.0 0.5 1.0 PCA 1 (98.6%) Figure 4 Relative growth of the length of sta- tolith dimensions. The relation be- tween the principal component score on the first axis for each .statolith against (A) statolith total length, and (B) age. The percentages indicate the variation in the data set that has been described by each principal component axis. Circles = adults; squares = juveniles. B Figure 5 Statoliths of Scfiiotcuthin IcuKimiana. demonstrating the change in shape from juvenile to adult squid. Mantle lengths and total statolith lengths were (A) 16 mm, 648 ^m; (B) 119 mm, 1.525 ;/m; (C) 190 mm, 1875 /jm, re- spectively. Statoliths are shown in an- terior view at different scales to facili- tate shape comparison. Table 3 Relative growth of statolith dimensions. Coefficient = coefficient of the first eigenvect ficient equals ( 1/p)" '' = 0.447. The difference equals the coefficient minus 0.447, which if allometric growth has occurred; data are logju-transformed; SE = standard errors. or. Isometric gi Tiust be greater owth than occurs when the coef- the critical difference Variable Coefficient Critical difference Difference Allometric growth Mean SE Total length 0.435 0.004 ± 0.008 -0.012 negative Dorsolateral 0.503 0.010 ± 0.022 0.056 positive Ventrolateral 0.387 0.006 ± 0.010 -0.060 negative Lateral dome 0.349 0.005 + 0.008 -0.098 negative Total width 0.535 0.007 + 0.017 0.088 positive fatter and bulkier during growth (Fig. 5). Statoliths of the adults continued to grow (Fig. 3); however, there is no relation between the PCA score on the first axis and statolith size for the adults (Fig. 4Ai, suggesting that the overall shape of individual adult statoliths showed no further change. Thomas and Moltschaniwskyi: Ontogenetic changes in size and shape of statoliths of Sepioleuth/s lessoniana 641 Table 4 Relative growth of statolith d imensions. Th e coefficients for the first three principal component axes (PCA)of the analysis on length variables; data are log,,, -transformed and the covariance matrix was used. Variable Eigenvector PCAl PCA2 PCA3 Total length 0.435 0.308 0.154 Dorsolateral 0.503 -0.695 0.321 Ventrolateral 0.387 0.592 -0.108 Lateral dome 0.349 0.210 0.579 Total width 0.535 -0.163 -0.726 Additional variation in statolith shape described on the second and third PCA axes was attributable to differences among juvenile, rather than adult squid (Fig. 6, A and B: Table 4). Growth of the sta- tolith along the ventrolateral and lateral dome axes and in total length were all important as statoliths approached their final size and shape. The ventro- lateral and total lengths were proportionally larger in juveniles, whereas the dorsolateral length and to- tal width were proportionally smaller immediately before individual statoliths attained their final adult shape. If the PCA score of each individual statolith on the first axis, as a descriptor of shape, is used to esti- mate age of the squid, then an improved correlation between age and statolith shape is seen when com- pared to the correlation between age and statolith total length alone (juveniles 77 = 15, 7-=0.73: adults 77=61, 7-'=0.61)( Fig. 4B). There is an exponential rela- tionship for both juvenile and adult squid but changes in statolith shape slow in older individuals. Analysis of adult and juvenile statoliths as separate gi'oups to eliminate gross statolith size differences that were iden- tified in the PCA only highlighted the overall variabil- ity in statolith shape within the two age groups, par- ticularly among adult squid. Discussion Results from this study suggest that neither statolith total length nor weight measurements can be reli- ably used to predict the age of a Sepioteuthis lessoniana individual. Statolith total length is a rea- sonable predictor of age for juvenile squid (less than 60 days old ) but cannot be used to predict age in older individuals of this species. This conclusion is in con- trast with that of Gonzalez etal. ( 1996) who observed that statolith length in ///ex coindetti shows signifi- 1.0- PCA 2(0.7%) ventro-lateral lateral dome all variables PCA 1(98.6? 1.0 dorso- .\ 0 !lateral PCA 3(0.3% )0-l ; lateral dome dorso-lateral --:'.« -0.1 dorso- lateral JSiL B PCA 2 •(0.7%) 0.1 ventro-lateral lateral dome ■0.1 jwidth ventro-lateral Figure 6 Relative growth of the length of sta- tolith dimensions. Principal component scores for the length of five statolith dimensions on the first three axes. The percentages indicate how much varia- tion in the data set has been described by each principal component axis. Data are logj„ transformed. Circles = adults; squares = juveniles. cant dependence on age (males 77 = 170, 7-^=0.71; fe- males 77 = 171, 7-=0. 74). However, they also suggested that the use of statolith length for age estimations would require verification over several years to con- firm this relationship. Statolith length has also been shown to reflect age in Illex illecebrosus females (77=31, 7^'=0.96) (Morris and Aldrich, 1985), whereas statolith weight provided a better reflection of age (77=112, 7^^=0.88 ) than statolith total length in Todarodes angolensis (Villanueva, 1992). Even combining all five statolith dimensions to produce a description of sta- tolith shape in this study provided only marginally better age estimates for ■§. lessoniana individuals than statolith total length or weight measurements alone. Variation was evident in size between adult sta- toliths in this study because the heaviest statoliths did not necessarily possess the longest total lengths. Additionally, as much as half of the variation in sta- tolith total lengths was not attributable to age of the 642 Fishery Bulletin 97(3), 1999 individuals. This indicates a disassociation between statolith and somatic (mantle length) growth in S. lessoniana. The disassociation of otolith and somatic growth has been demonstrated in several fish spe- cies (e.g. Mosegaard et al., 1988; Secor and Dean, 1989; Francis et al., 1993). In squids, Loligo chiiiensis (Jackson, 1995 ), and Todaropsis eblanae and Todarodes angolensis (Lipinski et al., 1993) also display disasso- ciation between statolith and somatic tissue growth. Comparison of statolith size and shape between Todarodes angolensis and Todaropsis eblanae also has revealed distinct species differences (Lipinski et al., 1993). Jackson (1995) suggested that the use of the relation of statolith length to mantle length as a pre- dictor of squid ages should proceed with caution until temperature-related effects on squid growth (e.g. sea- sonal effects), as well as the relation between statolith and somatic growth are better understood. The disassociation between statolith and somatic growth rates is possibly a function of differences be- tween the mechanisms responsible for these two pro- cesses. A process related to metabolic rate, rather than somatic growth, seems to govern the rate of otolith accretion in fish (Wright, 1991). Although a close correlation often exists between fish somatic growth and metabolic rate in early life-history stages, intrinsic or extrinsic constraints on somatic growth may affect this relationship and result in a disasso- ciation between otolith and somatic growth (Wright, 1991). Recently, Lombarte and Lleonart (1993) pro- posed that otolith growth in fish may occur under dual regulation: overall shape is genetically deter- mined whereas otolith size is governed by environ- mental factors. Several workers have also suggested that temperature plays a particularly important role in determining otolith growth, primarily through the effect of temperature on metabolic rate (Wright, 1991; Bradford and Geen, 1992). In order to demonstrate a link between otolith and somatic growth rates the age-independent variability in the relationship needs to be assessed (Hare and Cowen, 1995). The mecha- nisms governing statolith growth in S. lessoniana remain to be determined but may possibly be related to the high plasticity of somatic growth and a fixed growth trajectory of the statolith. Ontogenetic be- havioral adaptations may also play a key role. None- theless, there are important considerations for age and growth studies based on the morphological fea- tures of statoliths; whereas many squid are known to grow continuously throughout their life (Jackson and Choat, 1992; Jackson, 1994), growth ofstatoliths in our study appears to approach a final asymptotic size and shape. Statoliths grow with the age of the squid, but statolith accretion may respond differently to environ- mental factors than to growth of somatic tissue. The breakdown in the relation between statolith total length and age in older S. lessoniana individu- als may be attributed to variable accretion rates in the statoliths of adult squid, possibly in relation to environmental conditions. Varying aragonite accre- tion within statoliths could bring about fluctuations in daily increment widths that ultimately lead to differences in overall length and weight of similar aged statoliths. These differences are likely to be more detectable in the older, larger statoliths of adult squid. Alterations in otolith increment widths lead- ing to different sized otoliths of the same age have been shown for several fish species in response to both biotic and abiotic factors (e.g. Eckmann and Rey, 1987;Sogard, 1991; Burke etal., 1993). Further work on other fish species has shown that changes in in- crement width may lag or be unrelated to changes in somatic growth (Molony and Choat, 1990; Milicich and Choat, 1992). Variable statolith increment widths, that may be attributable to environmental conditions, have been shown in at least one squid species (Abralia trigonura, Bigelow, 1992), and one cuttlefish (Sepia hierredda, Raya et al. 1994). Modifications of otolith shape and daily increment structure in fish have been attributed to environmen- tal changes or fiuctuations in individual fish physi- ology (Morales-Nin, 1987; Nishimura, 1993; Wright, 1993; Tzeng and Tsai, 1994). As some fish migrate, modifications are often necessary in order to meet unique requirements for balance, orientation, and navigation ( Blaxter, 1988 ). A transformation of otolith shape may occur, resulting in discontinuities and secondary growth structures (Sogard, 1991; Hare and Cowen, 1994). Statoliths are the major sensory struc- ture responsible for balance and orientation of squid (Budelmann, 1990). The shape and structure of the statocyst chamber itself are important to the swim- ming performance and sensory perception capabili- ties of cephalopods (Williamson, 1991). Statolith shape and structure is, therefore, also likely to af- fect the response of this unique and complex sensory organ. Although S. lessoniana does not display dis- tinct life-style changes or habitat shifts, a transition from the observed juvenile habitat, where it com- monly shelters among floating surface debris, to an adult lifestyle that is typically more reef-associated, may involve behavioral adaptations that lead to dif- ferential growth of statoliths. Knowledge of the mechanism by which this growth occurs and how it can be modified during ontogeny will be critical to understanding how statoliths grow. Most growth of adult statoliths in this study oc- curred in the dorsal and lateral dome regions, pro- ducing a more rounded and bulkier form than that of the juveniles. The negative allometric growth in Thomas and Moltschaniwskyi: Ontogenetic changes in size and shape of statoliths of Sepioteuthis lessoniona 643 total length and along the lateral dome and ventro- lateral length axes suggests that relative growth of these regions is slow and that statoliths retain the elongate region of the rostrum as they grow. This was reflected in the superior definition and clarity of daily increment structures in the lateral dome re- gion of adult statoliths when viewed under a light microscope. This may also explain why optical acu- ity of daily increment structures in the rostrum re- gion of adult statoliths was poor. Slower growth of statoliths in the rostrum indicates deposition of less aragonite material compared with other faster grow- ing regions of the statolith and in turn suggests that daily increment widths in the rostrum will be nar- rower and therefore harder to discern with light mi- croscopy. In contrast, daily growth increments have been reported as displaying optimal definition in the rostrum of two other squid species (Photololigo edulis, Natsukari et al., 1988; Berryteuthis magister, Natsukari et al., 1993). Therefore, it is likely that differences occur in statolith accretion between spe- cies, possibly related to behavior. Optical definition of daily increments in statoliths has important con- sequences when ageing and growth studies are based on counts and distances between increments. Age and growth calculations based on narrower less distinct increments may not reveal true patterns otherwise evident in optically more distinct incrementation. Results from this study suggest that prediction of age of S. lessoniana individuals from the size and shape of their statoliths remains doubtful on account of the disassociation between statolith and somatic tissue growth. Changes in statolith shape during growth may also be important if statoliths are to be employed as a tool for constructing growth histories of S. lessoniana. Variable levels of statolith accre- tion ultimately leading to slight modifications in shape may not necessarily be indicative of changes in somatic growth. Ageing studies based on analyses of statolith size from other squid species need to be aware of the possible disassociation between statolith and somatic tissue growth. The changes in size and shape of Sepioteuthis lessoniona statoliths described in this study are likely due to differing aragonite accretion levels, possibly on a daily basis, in response to physi- ological and behavioral adaptations during ontogeny. Acknowledgments We thank G. Jackson for providing constructive re- views of early manuscripts. We also acknowledge J. Semniens, G. Peel, and P. Martinez for their assis- tance in the field and laboratory. This research was completed in partial fulfillment of an Honours de- gree by RT and was supported by a Merit Research Grant from James Cook University awarded to NAM. Literature cited Arkhipkin, A. 1993. Statolith microstructure and maximum age oiLoligo gain (Myopsida; Loliginidae) on the Patagonian Shelf. J. Mar. Biol. Assoc. U.K. 73:979-982. Bigclow, K. A. 1992. Age and growth in paralarvae of the mesopelagic squid A6ra/ia trigonura based on daily growth increments in statoliths. Mar Ecol. Prog. Ser. 82:.31-40. 1994. Age and growth of the oceanic squid Onychoteuthis bofcalijaponica in the North Pacific. Fish. Bull. 92:13-25. Blaster, J. H. S. 1988. Sensory performance, behaviour and ecology offish. In J. Atema, R. R. Fay, A. N. Popper, and W. N. Tauolga (eds.). Sensory biology of aquatic animals. Springer- Verlag, New York, NY. p. 203-232. Boehlert, G. W. 1985. Using objective criteria and multiple regression mod- els for age determination in fish. Fish. Bull. 83:103-117. Bradford, M. J., and G. H. Geen. 1992. Growth estimates from otolith increment widths of juvenile chinook salmon {Oncorhynch us tshawytscha) reared in changing environments. J. Fish Biol. 41:825-832. Budelmann, B. U. 1990. The statocysts of squid. In D. L. Gilbert. W. J. Adelman Jr., and J. M. Arnold (eds.). Squid as experimen- tal animals, p. 421-436. Plenum Press, London. Burke, J. S., D. S. Peters, and P. J. Hanson. 1993. Morphological indices and otolith microstructure of Atlantic croaker, Micropogonias undulatus. as indicators of habitat quality along an estuarine pollution gradient. Environ. Biol. Fishes 36:25-33. Campana, S. E., and J. D. Neilson. 1985. Microstructure of fish otoliths. Can. J. Fish. Aquat. Sci. 42:1014-1032. Clarke, M. R. 1978. The cephalopod statolith — an introduction to its form. J. Mar Biol. Assoc. U.K. 58:701-712. Eckmann, R., and P. Rey. 1987. Daily increments on the otoliths of larval and juve- nile Coregonus spp., and their modification by environmen- tal factors. Hydrobiologia 148:137-143. Fletcher, W. J. 1991. A test of the relationship between otolith weight and age for the pilchard Sardtnops neopilchardus. Can. J. Fish. Aquat. Sci. 48:3.5-38. Forsythe, J. W. 1993. A working hypothesis of how seasonal temperature change may impact the field growth of young cephalopods. In T. Okutani, R. K. O'Dor, and T. Kubodera (eds.). Recent advances in cephalopod fisheries biology, p. 133-143. Tokai Univ. Press. Tokyo. Porsythe, J. W., and R. T. Hanlon. 1989. Growth of the eastern Atlantic squid, Loligo forbesi Steenstrup (Mollusca: Cephalopoda I. Aquacult. Fish. Manage. 20:1-14. Francis, M. P., M. W. Williams, A. C. Pryce, S. Pollard, and S. G. Scott. 1993. Uncoupling of otolith and somatic growth in Pagriis aiiratus (Sparidael. Fish. Bull. 91:159-164. 644 Fishery Bulletin 97(3), 1999 Gonzalez, A. F., B. G. Castro, and A. Guerra. 1996. Age and growth of the short-finned squid Illex coindetti in Gahcian waters (NW Spain) based on statoHth analysis. ICES J. Mar. Sci. 53:802-810. Hare, J. A., and R. K. Cowen. 1994. Ontogeny and otolith microstructure of bluefish Pomatomus saltatrix (Pisces; Pomatomidaei. Mar. Biol. 118:541-550. 1995. Effect of age, growth rate, and ontogeny on the otolith size-fish size relationship in bluefish. Pomatomus saltatrix. and the implications for back-calculation of size in fish early life history stages. Can. J. Fish. Aquat. Sci. 52:1909-1922. Jackson, G.D. 1990. Age and growth of the tropical near shore loliginid squid Sepioteuthis lessoniana determined from statolith growth ring analysis. Fish. Bull. 88:113-118. 1994. Application and future potential of statolith incre- ment analysis in squids and sepioids. Can. J. Fish. Aquat. Sci. 51(1 1):2602-2626. 1995. Seasonal influences on statolith growth in the tropi- cal nearshore loliginid squid Loligo c/i;>it'/!S!s (Cephalopoda: Loliginidae) off Townsville, North Queensland, Australia. Fish. Bull. 93:749-752. Jackson, G. D., and J. H. Choat. 1992. Growth in tropical cephalopods: an analysis based on statolith microstructure. Can. J. Fish. Aquat. Sci. 49:218-228. Jolicoeur, P. 1963. The multivariate generalisation of the allometry equation. Biometrics 19:497-499. Jones, C. 1986. Determining age of larval fish with the otolith incre- ment technique. Fish. Bull. 84:91-103. Lipinski, M. R. 1986. Methods for the validation of squid age from statoliths. J. Mar. Biol. Assoc. U.K. 66:505-526. Lipinski, M. R. 1993. The deposition of statolith — a working hypothesis. In T. Okutani, R. K. O'Dor, and T. Kubodera (eds.). Recent advances in cephalopod fisheries biology, p. 241-262. Tokai Univ. Press, Tokyo. Lipinski, M. R., M. A. Compagno Roeleveld, and L. G. Underhill. 1993. Comparison of the statoliths of Todaropsis eblanaf and Todarodes angolensis (Cephalopoda: Ommastrephidae) in South African waters. In T. Okutani. R. K. O'Dor, and T. Kubodera (eds. I, Recent advances in cephalopod fisher- ies biology. Tokai Univ. Press, Tokyo, p. 263-273. Lombarte, A., and J. Lleonart. 1993. Otolith size changes related with body growth, habitat depth and temperature. Environ. Biol. Fishes 37:297-306. Marcus, L. 1990. Traditional morphometries. In F. J. Rohlf and F. L. Bookstein (eds.). Proceedings of the Michigan morpho- metries workshop, p. 77-122. Univ. Michigan Museum of Zoology, Special Publication 2 Milicich, M. J., and J. H. Choat. 1992. Do otoliths record changes in somatic growth rate? Conflicting evidence from a laboratory and field study of a temperate reef fish, Parika scaber. Aust. J. Mar. Fresh- water Res. 43:1203-1214. Molony, B. W., and J. H. Choat. 1990. Otolith increment widths and somatic growth rate: the presence of a time-lag. J. Fish Biol. 37:541-551. Morales-Nin, B. Y. O. 1987. The influence of environmental factors on microstruc- ture of otoliths of three demersal fish species caught off Namibia. S. Afr J. Mar. Sci. 5:255-262, Morris, C. C, and F. A. Aldrich. 1985. Statolith length and increment number for age de- termination o{ Illex illecebrosus (LeSueur. 1821) (Cephalo- poda: Ommastrephidae). NAFO Scientific Council Stud- ies 9:101-106. Mosegaard, H., H. Svcdang, and K. Taberman. 1988. Uncoupling of somatic and otolith growth rates in Arctic char tSalvelmus alpinus) as an effect of differences in temperature response. Can. J. Fish. Aquat. Sci. 45:1514-1524. Mugiya, Y., and S. Tanaka. 1992. Otolith development, increment formation, and an uncoupling of otolith to somatic growth rates in larval and juvenile goldfish. Nippon Suisan Gakkaishi. 58:845-851. Natsukari, Y. T., T. Nakanose, and K. Oda. 1988. Age and growth of loliginid squid Photololigo edulis (Hoyle, 1855). -J. Exp. Mar. Biol. Ecol. 116:177-190. Natsukari, Y., H. Mukai, S. Nakahama, and T. Kubodera. 1993. Age and growth estimation of a gonatid squid, Benyteuthis magister, based on statolith microstructure (Cephalopoda: Gonatidae). In T. Okutani, R. K. O'Dor, and T. Kubodera (eds. I, Recent advances in cephalopod fisher- ies biology, p. 351-364. Tokai Univ. Press, Tokyo. Nishimura, A. 1993. Occurrence of a check in otoliths of reared and sea- caught larval walleye pollock Theragra chalcogramma (Pallasi and its relationship to events in early-life history J. Exp. Mar. Biol. Ecol. 166:175-183. Radtke, R. L. 1983. Chemical and structural characteristics of statoliths from the short-finned squid Illex illecebrosus. Mar. Biol. 76:47-54. Raya, C. P., M. M. Fernandez-Nunez, E. Balguerias, and C. L. Hernandez-Gonzalez. 1994. Progress towards ageing cuttlefish {Sepia hierredda, Rang 1837 ) from North West African coast using statoliths. Mar Ecol. Prog. Ser. 114:139-147. Secor, D. H., and J. M. Dean. 1989. Somatic growth effects on the otolith-fish size rela- tionship in young pond-reared striped bass, Morone saxatilis. Can. J. Fish. Aquat. Sci. 46:113-121. Shea, B. T. 1985. Bivariate and multivariate growth allometry: statisti- cal and biological considerations. J. Zool. Lend. 206:367-390. Sogard, S. M. 1991. Interpretation of otolith microstructure in juvenile winter flounder iPseudopleuronectes americanus): ontoge- netic development, daily increment validation, and somatic growth relationships. Can. .J. Fish. Aquat. Sci. 48:1862- 1871. Stevenson, D. K., and S. E. Campana (eds.). 1992. Otolith microstructure examination and analysis. Can. Spec. Publ. Fish. Aquat. Sci. 117. 126 p. Tzeng, W. N., and Y. C. Tsai. 1994. Changes in otolith microchemistry of the Japanese ee\.Anguillajaponica. during its migration from the ocean to the rivers of Taiwan. .J. Fish Biol. 45:671-683. Villanueva, R. 1992. Interannual growth differences in the oceanic squid Todarodes angolensis Adam in the northern Benguela up- welling system, based on statolith growth increment analysis. J. Exp. Mar Biol. Ecol. 1,59:157-177. Williamson, R. 1991. Factors affecting the sensory respon.se characteris- Thomas and Moltschaniwskyi: Ontogenetic changes in size and shape of statoliths of Sepioteuthis lessoniana 645 tics of the cephalopod statocyst and their relevance in pre- 1993. Otohth microstructure of the lesser sandeel. Ammo- dieting swimming performance. Biol. Bull. 180:221-227. dytes marinus. J. Mar. Biol. Assoc. U.K. 73:245-248. Wright, P. J. Wright, P. J., N. B. Metcalfe, and J. E. Thorpe. 1991. The influence of metabolic rate on otolith increment 1990. Otolith and somatic growth rates in Atlantic salmon width in Atlantic salmon parr. Sa/mo.sa/arL. .J. Fish Biol. parr, Salmo salar L: evidence against coupling. J. Fish. 38:929-933. Biol. 36:241-249. 646 Abstract.— Economically valuable Cahfomiahahbut.Parahchthyscaliforn- iciis. and barred sand bass. Paralabrax nebuhfer, along with other fishes, are often abundant in the shallow areas of California bays during their early life history. However, little is known about their habitat use within these areas. We investigated habitat use of juvenile fishes in the shallow waters of an embayment by towing a 1.6-m beam trawl with 3-mm mesh through eelgrass beds [Zostefa manna) and unvegetated areas at depths <1.1 m in Alamitos Bay. Tows were conducted monthly or bi- monthly from May 1992 through No- vember 1995. A total of 435 tows dur- ing 31 months yielded 52,787 fishes comprising 46 species. However, the catch was dominated by only a few spe- cies and consisted mostly of juveniles and gobiid larvae. A total of 1157 Cali- fornia halibut and 225 barred sand bass were collected. California halibut were 2-6 times more abundant in unvegetated areas than in eelgrass beds, whereas barred sand bass were captured almost exclusively in eelgrass. Abundance of both species significantly decreased as distance from the bay mouth increased. Abundances of most other fishes also varied considerably between habitats and among sites. In contrast to Cali- fornia halibut and barred sand bass, abundances of other species were higher at sites farther inside the bay. Variations in water temperature, dis- solved o.\ygen, salinity, eelgrass shoot density, and eelgrass blade length failed to e.xplain differences in abundance for most fishes. Habitat and site selection for juvenile California halibut and barred sand bass may be related to lar- val supply and to the first suitable area encountered, but may be modified sub- sequently by movement into other ar- eas in search of preferred food items. Differential habitat use by California halibut, Paralichthys californicus, barred sand bass, Paralabrax nebulifer, and other juvenile fishes in Alamitos Bay, California Charles F. Valle John W. O'Brien Kris B. Wiese California Department of Fish and Game 330 Golden Shore, Suite 50 Long Beach, California 90802 E mail address (for C F Valle) cvalleSdfg2 cagov Manuscript accepted 25 June 1998. Fish. Bull. 97:646-660 ( 1999i. It is widely recognized that bays and estuaries are important nurs- ery grounds for many marine spe- cies. Within these areas, numerous studies have documented the impor- tance of eelgrass, Zostera marina, and other seagrasses as habitat for fishes. The composition and abun- dance of fishes in these habitats can vary considerably from unvegetated areas (Orth and Heck, 1980; Borton, 1982; reviewed in Orth et al., 1984; Heck et al., 1989; Ferrell and Bell, 1991; Sogard and Able, 1991). Seagrass habitats may be impor- tant because of their associated food resources or because they provide a refuge from predation (Adams, 1976; Heck and Thoman, 1981; re- viewed in Orth et al., 1984; Leber, 1985; Sogard and 011a, 1993). The association of fishes with seagrass beds has been related to various physical characteristics of seagrass, such as shoot density, blade length, and biomass (Adams, 1976; Orth and Heck, 1980; re- viewed in Orth et al., 1984; Bell and Westoby, 1986a). However, evidence suggests that physical characteris- tics of seagrasses may only affect fish abundances on a local scale such as within a seagrass bed, but not over larger scales such as dif- ferent beds within a bay (Bell and Westoby, 1986b; Bell et al., 1988; Sogard, 1989; Worthington et al.. 1992). Instead, it has been sug- gested that differing fish abun- dances in seagrass beds across an estuary are due to availability of competent larvae; pelagic larvae of some seagrass fishes settle indis- criminately in the first seagrass bed encountered regardless of seagrass physical characteristics (Bell and Westoby, 1986b; Bell et al., 1987, 1988). According to this "settle and stay" hypothesis, once within a seagrass bed, fishes would move around selecting microsites but would not leave the seagrass bed because of greater predation risks associated with moving over unvege- tated substrata. However, others have found that initial settlement patterns in habitats may be altered considerably by postsettlement mortality (Levin, 1994). migration to other areas (Sogard, 1989), or by both in response to available food (Jenkins et al., 1996). Much of the work on seagrasses and associated fishes has taken place on the east coast of the United States or in other parts of the world. Although several studies have de- scribed the ichthyofauna of south- ern California bays (Allen and Horn, 1975; Horn and Allen, 1976; Allen, 1982; Allen and Herbinson, 1991), few studies have described the relation of fishes with eelgrass and other habitats. Understanding Valle et aL: Habitat use by Paralichthys californicus and Paralabrax nebulifer 647 these relationships within bays is particularly im- portant owing to the destruction and severe alter- ation of about 75% of coastal estuary and wetland habitats in southern California since 1900 (Califor- nia Coastal Zone Conservation Commissions, 1975). Reduced catches of California halibut, Paralichthys californicus. may be due to the alteration or loss of this nursery habitat within bays and estuaries, or to both (Plummer et al., 1983; Kramer and Sunada, 1992; Kramer and Hunter^). California halibut is an important commercial and sport fish in southern and central California. Barred sand bass, Paralabrax nebulifer, is also an impor- tant sport fish in southern California and ranks an- nually among the top three species caught aboard commercial passenger fishing vessels (Love et al., 1996a). Both of these fishes spawn in nearshore wa- ters (Frey, 1971; Ono, 1992) and occupy embayments during their early life history; newly settled and larger juvenile California halibut are frequently found over shallow, sandy substrata (Haaker, 1975; Allen, 1988; Allen and Herbinson, 1990, 1991; Kramer, 1990, 1991a, 1991b), whereas juvenile barred sand bass are found in eelgrass beds (Feder et al., 1974; Rosales-Casian, 1997). However, little additional information is available on habitat use by these and other fishes within these areas. Therefore, the purpose of our study was to deter- mine 1) if abundances of juvenile California halibut, barred sand bass, and other fishes differed between eelgrass and unvegetated habitats, 2 ) whether these abundances differed among sites within the bay, and 3) whether these differences were related to physi- cal characteristics of eelgrass or abiotic factors. We examined habitat use by collecting fishes with a beam trawl towed in shallow eelgrass beds and nearby unvegetated areas at three sites within a single bay. A beam trawl was used because it collects smaller hali- but and other flatfishes more effectively than beach seines and otter trawls (Gunderson and Ellis, 1986; Kramer, 1990; Kuipers et al., 1992) and allows com- parison with other studies where similar gear was used. Materials and methods Study area and sampling Our study was conducted in Alamitos Bay (lat. 33°45'N, long. 118 = 07'W), which is a small embay- ment located at the southeastern boundary of Los Angeles County in southern California. Alamitos Bay was once an estuary of tidal marshes and mud flats. It has been considerably modified by dredging, fill- ing, and construction of homes, marinas, and two jetties that mark the entrance. The bay is exposed to semidiurnal tides with a mean range of 1. 1 m. Water circulation is further enhanced by large amounts of water drawn by two power stations that flush the bay every 19 hours (Phillips^). Regardless of tidal flux, there is a consistent flow of water into the bay (Brown and CaldwelP). Sampling was conducted at three sites (Bay En- trance, Belmont Shore, and Marine Stadium) sepa- rated by at least one km (Fig. 1). At each site, sam- pling occurred in two habitats, eelgrass (Zostera marina) beds and nearby unvegetated sandy-mud areas located about 40 m away. A weighted 1.6-m beam trawl, equipped with skis, tickler chain, and 3.0-mm stretched-mesh netting, was towed parallel to shore by two people on foot at low tide during the lowest tides of the month. Tows were made during daylight hours at depths from 0.3 m to 1.1 m, lasted 90 seconds, and covered approximately 56 m-. We completed 2-5 tows, depending on tide height, in each habitat at each site over four consecutive days. Sam- pling was conducted monthly from May 1992 through April 1993 (excluding February) and from Novem- ber 1993 through December 1994. Bimonthly sam- pling occurred from January 1995 through Novem- ber 1995. All fishes were sorted, identified, counted, and returned to the water. Most fishes were mea- sured to the nearest mm standard length (SL) from May 1992 through October 1994, whereas Califor- nia halibut and barred sand bass were measured throughout the study. Although California halibut and barred sand bass undergo transformation at about 7-9 mm SL and 11 mm SL, respectively (But- ler et al., 1982; Ahlstrom et al., 1984; Gadomski et al., 1990), all individuals <20 mm SL were consid- ered to be "newly settled" or "newly recruited" (Allen and Herbinson, 1990; Kramer, 1990; Love et al., 1996b). Larval and postlarval gobies (Brothers, 1975) that were not identified further were classified as "goby larvae." Other fishes not sexually mature based on size at first maturity information were considered to be juveniles. Water temperature, dissolved oxy- gen, and salinity data were collected for most tows, ' Kramer, S. H., and J. R. Hunter. 1987. Southern California wetlandyshallow water habitat investigation. Southwest Fish- eries Science Center, National Marine Fisheries Service, NOAA, P.O. Box 271, La Jolla, CA 92038. Ann. Rep. for Fiscal Year 1987, 12 p. ^ Phillips. K.W. 1978. A water and sediment study of Alamitos Bay correlated with peak and minimal recreational aquatic ac- tivities. City of Long Beach Chemical and Physical Testing Laboratory, Long Beach, California. Unpubl. data. ■^ Brown and Caldwell Environmental Sciences Division. 1979. Embaynient ecology studies: physical oceanographic and water quality data report. Southern California Edison, Unpubl. data. 648 Fishery Bulletin 97(3), 1999 500 meters Pacific Ocean Figure 1 Location of sampling sites (Bay Entrance, Belmont Shore, and Marine Stadium) in Alamitos Bay, California, U=unvegetated habitats, E=eelgrass habitats. except during September 1994 through December 1994, when dissolved oxygen values were not taken owing to equipment failure. Bottom water tempera- ture and dissolved oxygen were recorded with a YSI Model 5 IB oxygen meter. Surface salinity was mea- sured with a temperature-compensated refractometer. Eelgrass bed length, bed width, shoot density, and blade lengths were measured by two divers using SCUBA. A 300-m transect tape was laid out in the center of the bed along its longest axis. Width was measured perpendicular to the tape at three equi- distant points. Divers sampled 1/16 m^ quadrats in the bed at predetermined random points perpendicu- lar to the tape. Within each quadrat, all shoots were counted and the lengths of two randomly selected blades were measured. A total of 10-20 quadrats were sampled at each site during August 1992, March 1993, December 1993, and December 1995. Data analysis Nonparametric statistics were used because data and their transformations were heteroscedastic and not normally distributed. A nonparametric two-factor analysis of variance (ANOVA) on ranked data (Zar, 1984) was performed separately on the number of California halibut, barred sand bass, and 13 other common fishes captured per tow with habitat and site as factors. If significant main effects with no sig- nificant interactions were found, Tukey-Kramer multiple comparisons were performed to determine which pairs of means were significantly different. The results were considered significant if P was <0.05. If significant interactions between habitat and site were found, analyses of main effects were considered dubious and subsets of data were formed for each level of one factor within each level of the other fac- tor and vice-versa (Underwood, 1981). Kruskal-Wallis and Dunn multiple comparisons (Hollander and Wolfe, 1973) were performed on these subsets. For example, a Kruskal-Wallis test was used to deter- mine if there was an overall difference in abundance among sites for unvegetated habitats, and then for eelgrass habitats. If a significant difference was found, Dunn multiple comparisons were made to de- termine which pairs of sites were significantly dif- ferent. For fishes other than California halibut and barred sand bass, comparisons among sites were made only for the habitat where they were most abun- dant overall. Abundance differences between habitats were determined by a Wilcoxon two-sample test on unvegetated and eelgrass habitat data within each site. Two-sample Kolmogorov-Smirnov tests were used to compare length-frequency distributions between Valle et a\ Habitat use by Paralichthys califomicus and Paralabrax nebulifer 649 Table 1 Physical characteristics of eelgrass {Zostera manna) at three sites in Alaniitos Bay. Data were collected in August 1992, March 1993. December 1993, and December 1995. Values represent mean ± one standard error. Sample sizes for bed length and width measurements were 4 and 12, respectively. Sample sizes for density: Bay Entrance=79, Belmont Shore=70, and Marine Sta- dium=80. Sample sizes for blade length: Bay Entrance=154, Belmont Shore=138, and Marine Stadium=155. Bed length Site (m) Bed width Density (m) (no. shoots per quadrat) Blade length (cm) Bay Entrance 82.9 + 18.6 16.3 ±2.9 10.1 ±0.5 68.3 ±2.2 Belmont Shore 102.9 ± 2.5 8.8 ±0.3 7.9 ±0.5 37.2 ± 1.5 Marine Stadium 54.1 ± 5.9 15.8+1.3 10.6 ±0.7 47.9 ± 1.9 habitats and sites for California halibut and barred sand bass. Subsets of these data (see above) were used in these analyses. Kruskal-Wallis and Dunn multiple comparisons were also made on the mean number of eelgrass shoots and mean blade length per quadrat to deter- mine differences in these eelgrass characteristics among the three sites. Water temperature, dissolved oxygen, and salinity values for each tow were used in a two-way nonparametric ANOVA to test for dif- ferences in each of these abiotic factors between habi- tats and among sites. Results Eelgrass characteristics and abiotic factors Physical characteristics of eelgrass beds at the three sites were very different. Belmont Shore had the long- est bed but mean bed width, eelgrass shoot density, and eelgrass blade length were lowest (Table 1 ). Eel- grass densities at Bay Entrance and Marine Stadium were significantly greater than at Belmont Shore (P<0.05), and all blade length comparisons among sites were significantly different (P<0.05). Abiotic factors varied temporally but showed little spatial variation. Mean monthly values ranged from 14.5 to 23.0°C for water temperature, 4.8-8.1 mg/L for dissolved oxygen, and 23.2-36.0 ppt for salinity. These abiotic factors were very similar between habi- tats and among sites except for slightly lower dis- solved oxygen in eelgrass and higher water tempera- tures at Marine Stadium. However, none of these abiotic factors were significantly different between habitats or among sites (P>0.05). Fish community A total of 52,787 fish representing 46 species was collected from 435 tows (Table 2 ). The catch was domi- nated by a few species that were often captured in large numbers. Numbers of species varied among habitats and sites. Many more species were captured in eelgrass beds (/?=42) than in unvegetated areas (72=26). We collected 19 species exclusively in eelgrass beds but captured only three species solely in unvegetated areas (Table 2). Species numbers decreased as dis- tance from the bay mouth increased; more species were collected at Bay Entrance (n=40) than at Belmont Shore in =35) and Marine Stadium (n=28). California halibut California halibut ranked eighth in abundance; 1 157 individuals were collected from 50. 89^ of the tows (Table 2 ). The number of California halibut captured per tow ranged from 0 to 81 with a mean of 2.7 ± 0.32 SE. The abundance of newly settled Califor- nia halibut was greatest from March through May (Fig. 2). A total of 325 newly settled California hah- but was captured. For all sites and habitats com- bined, maximum mean density per month of newly settled individuals was 15/100 m- (May 1995). Maxi- mum mean density per month at an individual site and habitat was 81/100 m- (May 1995, Bay Entrance unvegetated area). California halibut abundance varied considerably among habitats and sites (Fig. 3A). The magnitude of these differences depended upon the habitat and site as indicated by a significant interaction term in the two-way ANOVA (P=0.002). Abundance was sig- nificantly different between habitats at all three sites, for all three site comparisons in unvegetated areas, and for two of three comparisons in eelgrass beds (Table 3). Within sites, Marine Stadium and Belmont Shore unvegetated areas contained about 2-3 times as many California halibut as eelgrass beds, and Bay Entrance unvegetated area had more than six times as many California halibut as nearby eelgrass (Fig. 3A). Within both habitats, California 650 Fishery Bulletin 97(3), 1999 Table 2 Total number of fishes captured by habitat type and frequency of occurrence in Alamitos Bay, California, from May 1992 through November 1995. Unvegetated Eelgrass Total Frequency of habitat habitat number occurrence C/f ) Common name Scientific name n=251 n=184 captured n=435 Goby larvae unidentified Gobiidae 6491 10336 16827 57.7 Cheekspot goby Ilypnus gilberti 8836 1009 9845 74.9 Bay pipefish Syngnathus leptorhynchus 70 8203 8273 49.0 Shiner perch Cymatogaster aggregata 5 4121 4126 23.2 Topsnielt Atherinops affinis 2463 1163 3626 33.1 Giant kelpfish Heterostichus rostratus 9 2454 2463 36.8 Arrow goby Clevelandia ios 1216 840 2056 18.6 California halibut Paralichthys californicus 1024 133 1157 50.8 Queenfish Seriphus politus 662 253 915 13.1 Diamond turbot Hypsopsetta guttulata 625 53 678 37.9 Spotted kelpfish Gibbonsia elegans 1 478 479 19.3 Shadow goby Quiet ula y-cauda 20 394 414 12.9 Pacific staghorn sculpin Leptocottus armatus 241 129 370 37.0 Bay blenny Hypsoblennius gentUis 3 326 329 20.5 Barred sand bass Paralabrax nebuhfer 4 221 225 12.4 Salema Xenistius californiensis 0 180 180 4.6 California killifish Fundulus parvipinnis 42 128 170 8.5 Black perch Embiotoca jackson i 0 86 86 5.3 Sargo Anisotremus davidsonii 0 86 86 3.7 Spotted turbot Pleuronichthys ritteri 54 12 66 8.7 Kelp bass Paralabrax clathratus 0 56 56 6.7 Snubnose pipefish Cosmocampus arctus 2 46 48 7.4 California clingfish Gobiesox rhessodon 4 41 45 5.1 Black croaker Cheilotrema saturnum 0 42 42 4.4 Mussel blenny Hypsoblennius jenkin si 7 35 42 3.9 Spotted sand bass Paralabrax maculatofasciatus 0 38 38 4.6 Reef finspot Paraclinus integripinnis 0 30 30 3.2 Dwarf surfperch Micrometrus minimus 0 28 28 2.5 Yellowfin goby Acanthogobius flavimanus 8 8 16 2.8 Mozambique tilapia Tilapia mossambica 0 14 14 2.1 Blenny larvae Hypsoblennius spp. 0 9 9 1.2 California tonguefish Symphurus atncauda 3 5 8 1.4 California scorpionfish Scorpaena guttata 0 6 6 1.2 Senorita Oxyjulis californica 0 4 4 0.9 White seabass Atractoscion nobilis 1 3 4 0.7 Bonefish Albula vulpes 2 2 4 0.5 Unidentified larvae unidentified Teleosti 4 0 4 0.5 Opaleye Girella nigricans 0 3 3 0.7 Rockpool blenny Hypsoblennius gilberti 0 3 3 0.2 Pacific sardine Sardmops sagax 2 0 2 0.5 Pile perch Rhacochilus vacca 0 2 2 0.5 Anchovy larva unidentified Engraulidae 1 0 0.2 Chocolate pipefish Syngnathus euchrous 0 1 0.2 Hornyhead turbot Pleuronichthys verticalis 0 1 0.2 Longjaw mudsucker Gillichthys mirabilis 0 1 0.2 Pacific barracuda Sphyraina argentea 1 0 0.2 Rock wrasse Halichoeres semicinctus 0 1 0.2 Striped mullet Mugil cephalus 1 0 0.2 White seaperch Phanerodon furcatus 0 1 0.2 Valle et aL: Habitat use by Paialichthys catifornicus and Paralabrax nebulifer 651 o o 16 7 Figure 2 Mean number per 100 m- of newly settled (<20 mm SLi and larger California halibut iParalichthys californicus) captured by month in Alamitos Bay from May 1992 through No- vember 1995 (n=435 towsl. Error bars represent + one standard error for the total number of halibut per 100 m- captured by month. halibut abundance decreased as distance from the bay mouth increased. Within unvegetated areas, Cahfomia hahbut were approximately 6 and 14 times more abundant at Bay Entrance than at Belmont Shore and Marine Stadium, respectively. California halibut length ranged from 7 to 253 mm SL, but few were larger than 140 mm SL; no adults were captured. Lengths were similar among habi- tats and sites except that many more newly settled individuals were captured in unvegetated areas than in eelgrass beds, particularly at Bay Entrance (Fig. 4). The length-frequency distribution of California halibut in unvegetated habitat at Bay Entrance was significantly different from nearby eelgrass (P=0.007 ) and from Belmont Shore unvegetated area (P=0.001 ). No other length-frequency comparisons were signifi- cantly different (P>0.05). Barred sand bass We collected 225 barred sand bass from 12.47f of the tows ( Table 2 ). The number of barred sand bass cap- tured per tow ranged from 0 to 42 with a mean of 0.5 ± 0.14 SE. Newly recruited barred sand bass and most larger juveniles were captured from Septem- ber through November (Fig. 5). Barred sand bass were captured almost exclusively (98.2% ) in eelgrass Table 3 Results of Kruskal-Wallis test iKW) and Dunn multiple comparisons (D) of California halibut iParalichthys californicus t abundance between habitats and among sites. df=degrees of freedom, H=Kruskal-Wallis statistic. BE=Bay Entrance, BS=Belmont Shore, MS=Marine Stadium, U=unvegetated habitat, and E=eelgrass iZostera marina) habitat. *=significant at P<0.0.5. Comparison Test df Statistic P-value Habitat (U vs. E) BE BS MS Site(U) BE vs. BS BE vs. MS BS vs. MS Site (El BE vs. BS BE vs. MS BS vs. MS KW 1 H=55.58* 0.0001 KW 1 //=16.03* 0.0001 KW 1 //=5.80* 0.0161 KW 2 //=115.23* 0.0001 D * <0.05 D * <0.05 D * <0.05 KW 2 //=22.65* 0.0001 D * <0.05 D * <0.05 D >0.05 beds and none were captured at Marine Stadium (Fig. 3B), leading to a significant interaction term in the two-way ANOVA (P=0.003). Because barred sand 652 Fishery Bulletin 97(3), 1999 bass were not captured at Marine Stadium and few were captured in unvegetated areas, only numbers within eelgrass beds at Bay Entrance and Belmont Shore were tested. Although approximately twice as many individuals were caught at Bay Entrance than at Belmont Shore, these results were not significantly different (P=0.68). Lengths of barred sand bass ranged from 16 to 92 mm SL but most (66.2%) were <40 mm SL; no adults ^ere captured. Average length at Bay Entrance (34.1 mm ±1.1 SE, n = 156) was smaller than at Belmont Shore (43.5 mm ±1.4 SE, n=67), primarily due to the greater abundance of new recruits and small juve- niles (21-30 mm SL) at Bay Entrance (Fig. 6). Figure 3 Mean number per 100 m- of (A) California hali- but iParalichthys californiciis) and (Bl barred sand bass iParalabrax nehulifer) captured in unvegetated and eelgrass iZostera marina) habi- tats at three sites in Alamitos Bay from May 1992 through November 1995. Dark bars represent unvegetated habitats and stippled bars represent eelgrass habitats. BE=Bay Entrance, BS=Bel- mont Shore, and MS=Manne Stadium. Number of tows in unvegetated habitats at BE, BS, and MS were 95, 61, and 95, respectively. Number of tows in eelgrass habitats at BE, BS, and MS were 60, 62, and 62, respectively. Error bars represent + one standard error. Length-frequency distributions of barred sand bass at these two sites were significantly different (P= 0.0001). Other species Gobies were the most abundant group of fishes, ac- counting for 55.2% of the total (Table 2). This group comprised mostly arrow goby (Clevelandia ios), cheekspot goby lllypnusgilberti), and numerous goby larvae that were probably arrow goby and cheekspot goby. Bay pipefish (Syngnathus leptorhynchus), shiner perch iCymatogaster aggregata), topsmelt (Atherinops affinis), and giant kelpfish (Heterostichus rostratus) were the other most abundant fishes, and along with gobies represented 87.9% of all individu- als collected. These fishes, along with queenfish (Seriphiis politus). diamond turbot (Hypsopsetta guttulata), spotted kelpfish (Gibbonsia elegans), shadow goby iQiiietula y-cauda). Pacific staghorn sculpin (Leptocottus armatus), and bay blenny (Hypsoblennius gentilis), were considered "common fishes." The common fishes, along with California halibut and barred sand bass, accounted for 96.5% of the total number collected. Abundances for most of these fishes peaked from March through May or from June through August; most of these fishes were juveniles. All of the common fishes were found in both habi- tats and at all sites; however the number of indi- viduals varied considerably (Fig. 7). Only queenfish and Pacific staghorn sculpin had no significant in- teractions between habitat and site in the nonpara- metric two-factor ANOVA (P>0.05). Queenfish were significantly more abundant in unvegetated areas than in eelgrass beds (P=0.0003). No significant habi- tat differences were found for Pacific staghorn sculpin (P=0.41). Queenfish were significantly more abun- dant at Marine Stadium than at Belmont Shore (P<0.05), whereas Pacific staghorn sculpin were sig- nificantly more abundant at Belmont Shore than at Bay Entrance (P<0.05). For 11 of the 13 common fishes, the nonparamet- ric two-way ANOVA yielded significant interactions between habitat and site (P<0.05 ). Stratifying by site and comparing abundances between eelgrass and unvegetated areas, we found that bay pipefish, shiner perch, giant kelpfish, spotted kelpfish, shadow goby, and bay blenny were significantly more abundant in eelgrass than in unvegetated areas at all three sites (P<0.05). Cheekspot goby and diamond turbot were significantly more abundant in unvegetated areas than eelgrass at all three sites, whereas goby larvae and arrow goby were significantly more abundant in unvegetated areas at Belmont Shore (P<0.05j. Valle et al : Habitat use by Paralichthys califomicus and Paralabrax nebulifer 653 ■O -o c o Z 280 240 200 20 1 BE (n=870) 0 20 40 60 80 100 120 140 160 180 200 220 240 260 0 20 40 60 80 100 120 140 160 180 200 220 240 260 MS (n=61) 0 20 40 60 80 100 120 140 160 180 200 220 240 260 24 20 16 12 0 6- 4- 10 4- BE (n=83) 0 20 40 60 80 too 120 140 160 180 200 220 240 260 10 8 oCH BS (n=29) 0 20 40 60 80 100 120 140 160 180 200 220 240 260 MS («=21) fcOo^ 0 20 40 60 80 too 120 140 160 180 200 220 240 260 Size class (mm) Figure 4 Length-frequency distributions of California halibut tParalichthys califomicus) captured in unvegetated and eelgrass iZostera manna ) habitats in Alamitos Bay from May 1992 through November 1995. Dark bars represent unvegetated habitats and stippled bars represent eelgrass habitats. See Figure 3 for further explanation. Topsmelt showed a mixed pattern; they were signifi- cantly more abundant in eelgrass at Marine Stadium (P<0.05l but significantly more abundant in unvegetated area at Belmont Shore (P<0.05). Stratifying by habitat and comparing abundances among sites, we found that most fishes were signifi- cantly more abundant at Marine Stadium or at Belmont Shore (P<0.05)( Table 4). Only shiner perch, giant kelpfish, and bay blenny had significantly more individuals at Bay Entrance than at one of the other two sites, but abundances at Bay Entrance were never significantly greater than at both of the other sites. Discussion California halibut and barred sand bass Abundance of California halibut was habitat specific. California halibut was one of the few fishes whose abundance was much higher in unvegetated areas than in eelgrass beds. Although eelgrass blades may provide shelter from predation for some inhabitants (Heck and Thoman, 1981; Sogard and OUa, 1993), California halibut typically avoid detection by preda- tors and prey by partially burying themselves in sedi- ment (Haaker, 1975). Other fiatfishes show substrate preferences (Tanda, 1990; Burke et al., 1991; Rogers, 1992 ), and California halibut <63 mm SL prefer bare sand over eelgrass in the laboratory (Drawbridge, 1990 ). Thus, sediments supporting eelgrass beds may not be preferable for settlement. It is also possible that eelgrass physically impedes halibut from set- tling there. If this were solely the case, then fewer halibut would be expected in eelgrass beds where dis- tances between shoots were shorter (dense beds I than in beds where distances between shoots were greater (sparse beds). However, we found this not to be true; more halibut were found in a dense eelgrass bed (Bay Entrance) than in a sparse bed (Belmont Shore). Barred sand bass abundance was also habitat spe- cific; they were almost exclusively found in eelgrass 654 Fishery Bulletin 97(3), 1999 ii 4- 73 UU > 20 mm ■1 <20mm X M M J J Month O D Figure 5 Mean number per 100 m- of newly recruited i<20 mm SLi and larger barred sand bass iParalabrax nebulifer) captured by month in Alamitos Bay from May 1992 through November 1995 (/!=435 tows). Error bars represent + one standard error for the total number of barred sand bass per 100 m' captured by month. 60 50 BE y. 3 4U > 30 C 20 -- ■7 in « 1 1=( 14 20 BS Ifi 12 8 4 1 ~nrn lu 20 .10 40 60 70 80 90 100 0 10 20 ,10 40 50 60 70 80 90 100 Size class (mm) Figure 6 Length-frequency distributions of barred sand bass iParalabrax nebulifer) captured in eel- grass (Zostera marina) habitat in Alamitos Bay from May 1992 through November 1995. See Figure 3 for further explanation beds. Eelgrass beds may be productive foraging areas for them. Although the diet of newly recruited and small juvenile barred sand bass is not known, larger juvenile barred sand bass (123-239 mm SL) consume amphi- pods, shrimps, and other small crustaceans (Roberts et al., 1984). These items are often abundant in eel- grass beds, and this plentiful food supply may en- able them to achieve faster growth rates, enabling them to achieve a size that is less vulnerable to pre- dation more quickly (Levin et al., 1997). Additionally, eelgrass may offer shelter because most newly recruited and juvenile barred sand bass have been observed around eelgrass or other structure such as mussels, rocks, or debris during SCUBA surveys (senior author's unpubl. data). Larger juveniles and adults are mostly found over sandy bottoms and among rocks (Turner et al., 1969; Feder et al.,1974) and are scarce in eelgrass beds (senior author's unpubl. data). Valle et al.: Habitat use by Paralichthys ca/ifornicus and Paralabrax nebulifer 655 Queenfish 0 .L5 - i Pacific T slaghorn ^| sculpin ^^k i Ih rr 1 50 30 Gia nt k T elpfis 1 T 20 10 r^ Spotted kelpfish 1 .r!i_ ri 300 Bay pipefish in Shadow goby J^L Goby lar\'ae BE BS ii MS ^u Shiner perch r 90 60 30 -- l""l Bay blenny T T Arrow goby ^ BE BS MS Figure 7 Mean number per 100 m- of the most abundant fishes (see text) captured in unvegetated and eelgrass (Zostera marina) habitats at three sites in Alamitos Bay from May 1992 through November 1995. Dark bars represent unvegetated habitats and stippled bars represent eelgrass habitats. See Figure 3 for further explanation. The distribution patterns of California halibut and barred sand bass among sites were similar; abun- dances of individuals decreased with increasing dis- tance from the bay mouth. These patterns in Alamitos Bay might be expected if abundances were related to larval supply because both species spawn outside the bay. Others have found that circulation patterns may limit dispersal of recruits to inner parts of an embaymentlSogardetal., 1987; Jenkins etal., 1996). However, this does not seem to be the case for Alamitos Bay because it is well circulated with a net inflow of water, and the distance between sites is relatively short. Instead, fewer halibut and barred sand bass may inhabit the inner parts of the bay because settlement and location of suitable habitat have occurred before reaching these areas. For Cali- fornia halibut, few larvae are found in embayments (White. 1977; Leithiser, 1981; Nordby, 1982; Yokla- vich et al., 1992); the greatest densities of eggs and early larvae occur in nearshore waters with older larvae found less than 1 km from shore (Ahlstrom and Moser, 1975; Gruber et al., 1982; Barnett et al., 1984; Lavenberg et al., 1986; Walker et al., 1987; Moser and Watson, 1990). Most transforming hali- but larvae are found on the open coast (Kramer, 1990, 1991a), and large numbers of halibut settle there and in embayments (Allen, 1988; Allen and Herbinson, 1990, 1991; Kramer, 1990. 1991a, 1991b). Those hali- but that settle on the open coast move into embay- ments or die (Kramer, 1991a). Although it is not known where most barred sand bass recruitment occurs, spawning occurs in nearshore waters, and the planktonic larval phase is relatively short, lasting about one month (Butler et al., 1982). Therefore, if 656 Fishery Bulletin 97(3), 1999 Table 4 Results of Dunn mu Itiple com parisons of fish abundance among sites BE=Bay Entrance, BS=Belmont Shore and MS=Marine Stadium. Comparisons among sites were done only for the habitat (eelgrass or unvegetated ) where fishes were most abundant. * indicates which site showed a significantly greater abund anceatP<0.05. Common name Scientific name BE vs. BS BE vs. MS BS vs. MS Eelgrass Bay pipefish Syngnathus leptorhynchus *MS *MS Shiner perch Cymatogaster aggregata *BE *MS ■^ Giant kelpfish Heterostichus rostratus *BS *BE *BS Spotted kelpfish Gibbonsia elegans *BS *BS Shadow goby Quietula y-cauda *MS Bay blenny Hypsoblennius gentilis *BE *BS Goby larvae Gobiidae *MS *MS Unvegetated Cheekspot goby Ilypnus gilberti *BS *MS Diamond turbot Hypsopsetta guttulata *BS *MS *MS Arrow goby Clevelandia ios *BS *MS Topsmelt Atherinops affinis *BS habitat is suitable, one would expect more newly settled California halibut and newly recruited barred sand bass at sites nearest the entrance because they are more likely to encounter these areas first. This might be especially true for embayments where the entrance is small, such as in Alamitos Bay, which could enhance chances of individuals encountering a shallow site nearest the mouth of the bay. It is also possible that the distribution patterns we observed were due to active site selection. Dia- mond turbot, whose larvae also occur in nearshore waters (Barnettetal., 1984; Walker et al., 1987) and are thus exposed to the same hydrodynamic pro- cesses, were most abundant in the inner part of the bay and least abundant nearest the mouth of the bay. Kramer (1991b) also obtained opposite distribution patterns for juvenile California halibut and diamond turbot. Burke (1995) found that the distribution of two newly settled flounders that immigrated to an estuary were influenced by the availability of pre- ferred prey types. Although we have no data on in- vertebrate prey distributions among sites, it is pos- sible that California halibut and diamond turbot also select sites on the basis of availability of their pre- ferred food items. Juvenile California halibut and diamond turbot have very different diets: small ju- venile California halibut feed mainly on small crus- taceans (Haaker, 1975; Plummer et al., 1983; Allen, 1988; Drawbridge, 1990), whereas juvenile diamond turbot feed mainly on polychaetes (Lane, 1975). As halibut grow larger, they become more piscivorous and gobies become an increasingly important part of their diet (Allen, 1988; Drawbridge, 1990). This might explain why we found relatively larger halibut at Belmont Shore than at Bay Entrance because this area contained more arrow gobies and cheekspot gobies. Inhabiting an area with preferred food items might lead to accelerated growth rates, which may be an advantage for reducing size-selective preda- tion. Sogard (1992) found that winter flounder iPleuronectes americanus) and tautog (Tautoga onitis) were most abundant in estuarine areas that supported faster growth rates. However, we can not discount that larval supply may also be a factor be- cause the spawning locations of diamond turbot are not known. Lane (1975) suggested that spawning takes place in or very near the outer harbor. This area would be much closer than where halibut spawn, and inner parts of the bay would be closer to eggs and larvae. In addition, laboratory experiments in- dicate that diamond turbot larvae are able to sur- vive longer periods of starvation than halibut larvae (Gadomski and Petersen, 1988). This may enable diamond turbot larvae to remain in the water col- umn longer and reach inner parts of the bay whereas halibut larvae must settle earlier Barred sand bass may also actively select sites within Alamitos Bay. However, the differences in physical characteristics of eelgrass among sites ap- peared not to affect the abundance of new recruits and juveniles. Although Bay Entrance had signifi- cantly more eelgrass shoots and longer blade lengths than Belmont Shore, barred sand bass abundances at the two sites were not significantly different. This lack of difference suggests that the effect of seagrass physical characteristics on fish abundance breaks down over larger spatial scales. However, if barred sand bass settle indiscriminately into the first eelgrass bed encountered, we would have expected significantly more individuals nearest the bay mouth. Sogard ( 1989) Valle et al.: Habitat use by Paralichthys cahfornicus and Paralabrax nebulifer 657 found that initial settlement patterns for some fishes in seagrass beds were altered considerably by move- ment to other areas. Our results indicate that this may be also true for barred sand bass because significantly larger fish were found at Belmont Shore, indicating that movement to this site from the site nearest the entrance after initial settlement may have occurred. However, we are not able to resolve why barred sand bass were absent from the inner part of the bay. Densities of newly settled and juvenile California halibut in our study were much greater than those reported by others using similar gear in other bays. Kramer (1990) obtained maximum monthly means for shoreline habitats in Agua Hedionda Lagoon of 3.7 newly settled California halibut per 100 m-^ for all stations combined and 9.2 per 100 m' for a single station. These densities are approximately 25'7f and 11"^, respectively, of the values we obtained for Alamitos Bay. Our maximum monthly mean num- ber of newly settled California halibut was approxi- mately four times greater than that obtained for nearby Anaheim Bay in 1989 (Allen and Herbinson, 1990). Thus, our results support the conclusion by Allen and Herbinson ( 1990) that there is a great deal of annual variability in numbers of newly settled and juvenile California halibut within and among embayments in southern California. Although the beam trawl may capture demersal fishes such as California halibut less effectively in eelgrass beds than in unvegetated areas, this seems an unlikely explanation for the great differences in habitat specific fish abundances that we observed. First, we collected large amounts of eelgrass blades (occasionally with attached rhizome) in many tows, and SCUBA observations indicated that the beam trawl maintained continuous contact with the sub- strate in both habitats. Second, staghorn sculpin, which rest on the substrate, and shadow goby, which often burrow in the sediment, were frequently cap- tured in both habitats. Because we collected data only at low tides during daylight hours, we have no infor- mation on how sampling at different tide heights and times of the day would have affected our results. Diel differences in fish abundance can occur in southern California bays (Horn, 1980; Hoffman^); however, there are no data to suggest that these differences are related to habitats or locations within the bay. Other species Alamitos Bay was typical of other temperate bay- estuarine environments in having relatively few spe- cies account for a large proportion of the total num- ber of individuals collected (reviewed in Allen and Horn, 1975: Horn, 1980; Allen, 1982; Onuf and Quam- men, 1983; Allen and Herbinson, 1991; Hoffman-*; MBC Applied Environmental Sciences'^). Species composition and abundance were very similar to data collected us- ing similar gear in Anaheim Bay and Agua Hedionda Lagoon (Allen and Herbinson, 1991; MBC Applied En- vironmental Sciences''). Our results, however, lacked high abundances of northern anchovy (Engraulis nwrdox), slough anchovy (Anchoa delicatissima), deepbody anchovy (Anchoa compressa ), and California killifish iFuiidulus pafvipinnis) found by other stud- ies in the inner part of Alamitos Bay and Newport Bay (Allen and Horn, 1975; Allen, 1982). Differences in sam- pling gear and location are probably responsible, but other factors may also play a role. As noted by Allen and Horn ( 1975), the abundance of northern anchovy may have been due to high periods of recruitment. In- deed, recruitment biomass of northern anchovy (age zero at 1 July ) off southern California during their stud- ies (1973 and 1979) was 2.6-3.1 times greater than during 1992-94 ( Jacobson et al.*^) For most species, greater abundances of individu- als in either eelgrass or unvegetated areas suggested habitat preferences or differential mortality for these species in shallow water of Alamitos Bay. The large numbers of species and individuals in eelgrass beds indicated the importance of this habitat for many fishes, especially juveniles during spring and sum- mer. Many of the fishes also differed in abundance across sites. More species were found near the bay mouth owing in part, to the occasional presence of more typical nearshore residents, but more individu- als of several species were found farther inside the bay. However, the differences in eelgrass physical characteristics among sites did not appear to affect the abundances of most fishes. For example, abun- dances of bay blenny, bay pipefish, and shadow goby were not significantly different between Bay En- trance and Belmont Shore, although eelgrass den- sity and blade lengths were significantly different between these sites. In addition, giant kelpfish and spotted kelpfish were most abundant at Belmont Shore, the site with the lowest eelgrass density and '' Hoffman, R. S. 1986. Fishery utilization of eelgrass (Zos^era marina) beds and non-vegetated shallow water areas in San Diego Bay. National Marine Fisheries .Service, Southwest Re- gion. .50"l W. Ocean Blvd. Suite 4200. Long Beach. CA. 90802. Admin. Rep. SWR-86-4, 29 p. ^ MBC Applied Environmental Sciences. 1990. Distribution of ju- venile California halibut iPaj-alichthys californicus) and other fishes in bay and coastal habitats of Los Angeles, Orange, and San Diego counties in 1989. MBC Applied Environmental Sci- ences, 947 Newhall Street. Costa Mesa, CA 92627. Final Rep. 90-RD-09. 27 p. *^ Jacobson. L. D.. N. C. H. Lo. S. F Hemck Jr. and T. Bishop. 1995. Spawning biomass of the northern anchovy in 1995 and status of the coastal pelagic fishery during 1994. Southwest Fisheries Science Center. National Marine Fisheries Service. NOAA, P. O. Box 271, La Jolla, CA 92038. Admin. Rep. U-95-11, 49 p. 658 Fishery Bulletin 97(3), 1999 shortest blade lengths. Only shiner perch abundance appeared related to eelgrass physical characteristics. This was not surprising because shiner perch are closely associated with the amount of eelgrass cover (Onuf and Quammen, 1983). In conclusion, we found that for Alamitos Bay: 1) shallow unvegetated and eelgrass habitats were im- portant for many fishes, especially juveniles, 2) ju- venile California halibut and barred sand bass used different habitats; California halibut inhabited unvegetated areas and barred sand bass inhabited eelgrass beds, 3 ) habitats nearest the bay mouth were particularly important for juvenile California hali- but and barred sand bass, whereas habitats farther inside the bay were more important for other fishes, 4) habitat and site selection for juvenile California halibut, barred sand bass, and most other fishes ap- peared unrelated to physical characteristics of eel- grass or abiotic factors, 5) habitat and site selection for juvenile California halibut and barred sand bass may be related to larval supply and to the first suit- able habitat and site encountered, but may be modi- fied subsequently by movement into other areas in search of preferred food items. A closer look at shal- low unvegetated and eelgrass habitats in other bays in relation to California halibut and barred sand bass abundance is warranted. Protection of these habitats from elimination or even alteration may be important for the successful management of these species. Acknowledgments This work was primarily funded through the Fed- eral Aid in Sport Fish Restoration Act. We thank the following individuals for invaluable assistance in the field: Mary Larson, Paul Gregory, Kris Eager, other California Dept. of Fish and Game staff, and students from California State University at Long Beach. Calvin Chun helped with the statistical analyses. We also thank David Parker, Rick Klingbeil, Larry Allen, Calvin Chun, and three anonymous reviewers who provided valuable comments on this manuscript. Literature cited Adams, S. M. 1976. The ecology of eelgrass, Zostera manna (L. I, fish com- munities. I. Structural analysis. J. Exp. Mar Biol. Ecol. 22:269-291. Ahlstrom, E. H., K. Amaoka, D. A. Hensley, H. G. Moser, and B. Y. Sumida. 1984. Pleuronectiformes:development. /;j H. G. Moser, W. .J Richards, D. M. Cohen, M. P. Fahay, A. W. Kendall Jr., and S. L. Richardson (eds.). Ontogeny and .systematies of fishes, p. 640-670. Am. Soc. Ichthyol. Herpetol., Spec. Publ. 1. Ahlstrom, E. H., and H. G. Moser. 1975. Distributional atlas offish larvae in the California Current region: flatfishes, 1955-1960. Calif Coop. Oce- anic Fish. Invest. Atlas 23. Allen, L. G. 1982. Seasonal abundance, composition, and productivity of the littoral fish assemblage in upper Newport Bay, California. Fish. Bull. 80:769-789. 1988. Recruitment, distribution, and feeding habits of young-of-the-year California halibut (Paralichthys califor- nicus) in the vicinity of Alamitos Bay-Long Beach Harbor, California, 1983-1985. Bull. So. Calif Acad. Sci. 87:19- 30. Allen, L. G., and M. H. Horn. 1975. Abundance, diversity, and seasonality of fishes in Colorado Lagoon, Alamitos Bay, California. Estuarine Coastal Mar Sci. 3:371-380. Allen, M. J., and K. T. Herbinson. 1990. Settlement of juvenile California halibut, Paralich- thys californicus, along the coasts of Los Angeles. Orange, and San Diego Counties in 1989. Calif Coop. Oceanic Fish. Invest. Rep. 31:84-96. 1991. Beam-trawl survey of bay and nearshore fishes of the soft-bottom habitat of southern California in 1989. Ca- lif Coop. Oceanic Fish. Invest. Rep. 32:112-127. Barnett, A. M., A. E. Jahn, P. D. Sertic, and W. Watson. 1984. Distribution of ichthyoplankton off San Onofre, Cali- fornia, and methods for sampling very shallow coastal waters. Fish. Bull. 82:97-111. Bell, J. D., A. S. Steffe, and M. Westoby. 1988. Location of seagrass beds in estuaries: effects on as- sociated fish and decapods. J. Exp. Mar. Biol. Ecol. 122:127-146. Bell, J. D., and M. Westoby. 1986a. Importance of local changes in leaf height and den- sity to fish and decapods associated with seagrasses. J. Exp. Mar Biol. Ecol. 104:249-274. 1986b. Variation in seagrass height and density over a wide spatial scale: effects on common fish and decapods. J. Exp. Mar Biol. Ecol. 104:27.5-295. Bell, J. D., M. Westoby, and A. S. Steffe. 1987. Fish larvae settling in seagrass: do they discriminate between beds of different leaf density? J, Exp. Mar Biol. Ecol. 111:133-144. Borton, S. F. 1982. A structural comparison of fish assemblages from eelgrass and sand habitats at Alki Point, Washington. M.S. thesis, Univ. Washington, Seattle, WA, 85 p. Brothers, E. B. 1975. Comparative ecology and behavior of three sympat- ric gobies. Ph.D. diss., L'niv. Calif, San Diego, CA. 370 p. Burke, J. S. 1995. Role of feeding and prey distribution of summer and southern flounder in selection of estuarine nursery habitats, J. Fish Biol. 47:35.5-366. Burke, J. S., J. M. Miller, and D. E. Hoss. 1991. Immigration and .settlement pattern ofParalichthys dentatus and P. lethostigma in an estuarine nursery ground. North Carolina, U.S.A. Neth. J. Sea Res. 27:393-405. Butler, J. L., H. G. Moser, G. S. Hageman, and L. E. Nordgren. 1982. Developmental stages of three California sea basses tParalnbrax. Pisces, Serranidae). Calif Coop. Oceanic Fish. Invest. Rep. 23:252-268. California Coastal Zone Conservation Commissions. 1975. California coastal plan. State Printing Office, Sac- ramento, CA, 443 p. Valle et al.: Habitat use by Paraltchthys califomicus and Paralabiax nebuliter 659 Drawbridge, M.A. 1990. Feeding relationships, feeding activity and substrate preferences of juvenile California halibut, Paralichthys califomicus, in coastal and bay habitats. M.S. thesis, San Diego State Univ, San Diego. CA, 214 p. Feder, H. M., C. H. Turner, and C. Limbaugh. 1974. Observations on fishes associated with kelp beds in southern California. Calif. Dep. Fish Game, Fish Bull. 160, 144 p. Ferrell, D. J., and J. D. Bell. 1991. Differences among assemblages of fish associated with Zostera capricorni and bare sand over a large spatial scale. Mar. Ecol. Prog. Ser. 72:15-24. Frey, W. H., ed. 1971. California's living marine resources and their utiliza- tion. Calif Dep. Fish Game, Sacramento, CA, 148 p. Gadomski, D. M., S. M. Caddell, L. R. Abbott, and T. C. Caro. 1990. Growth and development of larval and juvenile Cali- fornia halibut, Paralichthys califomicus, reared in the laboratory. In C. W. Haugen (ed.). The California hali- but, Paralichthys califomicus, resource and fisheries, p. 8.5-98. Calif Dep. Fish Game. Fish Bull. 174. Gadomski, D. M., and J .H. Petersen. 1988. Effects of food deprivation on larvae of two flatfishes. Mar. Ecol. Prog. Ser. 44:103-111. Gruber, D., E. H. Ahlstrom, and M. M. Mullin. 1982. Distribution of ichthyoplankton in the Southern Cali- fornia Bight. Calif Coop. Oceanic Fish. Invest. Rep. 23: 172-179. Gunderson, D. R., and I. E. Ellis. 1986. Development of a plumb staff beam trawl for sam- pling demersal fauna. Fish. Res. 4:35-41. Haaker, P. L. 1975. The biology of the California halibut, Paralichthys califomicus (Ayres). in Anaheim Bay, California. /;; E. D. Lane and C. W. Hill (eds.). The marine resources of Anaheim Bay p. 137-151. Calif Dep. Fish Game, Fish Bull. 165. Heck, K. L., Jr., K. W. Able, M. P. Fahay, and C. T. Roman. 1989. Fishes and decapod crustaceans of Cape Cod eelgrass meadows: species composition, seasonal abundance pat- terns and comparison with unvegetated substrates. Estu- aries 12:59-65. Heck, K. L., Jr., and T. A. Thoman. 1981. Experiments on predator-prey interactions in veg- etated aquatic habitats. J. Exp. Mar. Biol. Ecol. 53:125- 134. Hollander, M., and D. A. Wolfe. 1973. Nonparametric statistical methods. John Wiley and Sons, New York, NY, 503 p. Horn, M. H. 1980. Diel and seasonal variation in abundance and diver- sity of shallow-water fish populations in Morro Bay, California. Fi.sh. Bull. 78:759-770. Horn, M. H., and L. G. Allen. 1976. Numbers of species and faunal resemblance of ma- rine fishes in California bays and estuaries. Bull. So. Cal. Acad. Sci. 75:159-170. Jenkins, G. P., M. J. Wheatley, and A. G. B. Poore. 1996. Spatial variation in recruitment, growth, and feed- ing of postsettlement King George whiting, Sillagniodes punctata, associated with seagrass beds of Port Phillip Bay, Australia. Can. J. Fish. Aquat. Sci. 53:350-3.59. Kramer, S. H. 1990. Distribution and abundance of juvenile California halibut, Paralichthys californicus, in shallow waters of San Diego County. In C. W. Haugen (ed. ), The California hali- but, Paralichthys califomicus, resource and fisheries, p. 99-126. Calif Dep. Fish Game, Fish Bull. 174. 1991a. Growth, mortality, and movements of juvenile Cali- fornia halibut, Paralichthys califomicus in shallow coastal and bay habitats of San Diego County, California. Fish. Bull. 89:195-207. 1991b. The shallow-water flatfishes of San Diego County. Calif Coop. Oceanic Fish. Invest. Rep. 32:128-142. Kramer, S. H., and J. S. Sunada. 1992. California halibut. In W. Leet, C. M. Dewees, and C. W. Haugen (eds.), California's living marine resources and their utilization, p. 94-97. Univ. California Sea Grant Extension Program, Davis, CA. Kuipers, B. R., B. MacCurrin, J. M. Miller, H. W. van der Veer, and J. I. J. Witte. 1992. Small trawls in juvenile flatfish research: their de- velopment and efficiency. Neth. J. Sea Res. 29:109-117. Lane, E. D. 1975. Quantitative aspects of the life history of the diamond turbot, Hypsopsetta guttulata (Girard), in Anaheim Bay. In E. D. Lane and C. W. Hill (eds.). The marine resources of Anaheim Bay. p. 153-173. Calif Dep. Fish Game, Fish Bufl. 165. Lavenberg, R. J., G. E. McGowan, A. E. Jahn, J. H. Petersen, and T. C. Sciarrota. 1986. Abundance of southern California nearshore ichthyo- plankton: 1979-1984. Cahf Coop. Oceanic Fish. Invest. Rep. 27:53-64. Leber, K. M. 1985. The influence of predatory decapods, refuge, and microhabitat selection on seagrass communities. Ecology 66:1951-1964. Leithiser, R. M. 1981. Distribution and seasonal abundance of larval fishes in a pristine southern California salt marsh. Rapp. P.-V. Reuns. Cons, Int. Explor. Mer 178:174-175. Levin, P. S. 1994. Fine-scale temporal variation in recruitment of a tem- perate demersal fish: the importance of settlement versus post-settlement loss. Oecologia 97:124-133. Levin, P. S., R. Petrik, and J. Malone. 1997. Interactive effects of habitat selection, food supply and predatinn on recruitment of an estuarine fish. Oeco- logia 112:55-63. Love, M. S., A. Brooks, and J. R. R. Ally. 1996a. An analysis of commercial passenger fishing vessel fisheries for kelp bass and barred sand bass in the South- ern California Bight. Calif Fish and Game 82:105-121. Love, M. S., A. Brooks, D. Busatto, J. Stephens, and P. A. Gregory. 1996b. Aspects of the life histories of the kelp bass. Paralahrax clathratus. and barred sand bass, P. nebulifer, from the Southern California Bight. Fish. Bull. 94:472- 481. Moser, H. G., and W. Watson. 1990. Distribution and abundance of early life history stages of the California halibut, Paralichthys califomicus, and comparisons with the fantail sole, Xystreurys liolepis. In C. W. Haugen (ed.). The California halibut, Paralichthys californicus, resource and fisheries, p. 31- 84. Calif Dep. Fish Game, Fish Bull. 174. Nordby, C. S. 1982. The comparative ecology of ichthyoplankton within Tijuana Estuary and in adjacent nearshore waters. M.S. thesis, San Diego State Univ., San Diego, CA, 101 p. 660 Fishery Bulletin 97(3), 1999 Ono, D. S. 1992. Sand basses. In W. Leet, C. M. Dewees, and C. W. Haugen (eds), California's living marine resources and their utilization, p. 151-153. Univ. California Sea Grant Ex- tension Program, Davis, CA. Onuf, C. P., and M. L. Quammen. 1983. Fishes in a California coastal lagoon: effects of major storms on distribution and abundance. Mar. Ecol. Prog. Ser 12:1-14. Orth, R. J., and K. L. Heck Jr. 1980. Structural components of eelgrass (Zostera marina) meadows in the lower Chesapeake Bay-fishes. Estuaries ■^ 3:278-288. Orth, R. J., K. L. Heck Jr., and J. V. van Montrfans. 1984. Faunal communities in seagrass beds: a review of the influence of plant structure and prey characteristics on predator-prey relationships. Estuaries 7:339-350. Plummer, K. M., E. E. Demartini, and D. A. Roberts. 1983. The feeding habits and distribution of juvenile-small adult California halibut, (Paralichthys californicus) in coastal waters off northern San Diego county. Calif Coop. Oceanic Fish. Invest. Rep. 24:194-201. Roberts, D. A., E. E. DeMartini, and K. M. Plummer. 1984. The feeding habits of juvenile-small adult barred sand bass iParalabrax nebulifer) in nearshore waters off north- ern San Diego County. Calif Coop. Oceanic Fish. Invest. Rep. 25:105-111. Rogers, S. I. 1992. Environmental factors affecting the distribution of sole iSolea solea (L.)l within a nursery area. Neth. J. Sea Res. 29:153-161. Rosales-Casian, J. A. 1997. Inshore soft-bottom fishes of two coastal lagoons on the northern Pacific Coast of Baja California. Calif. Coop. Oceanic Fish. Invest. Rep. 38:180-192. Sogard, S. M. 1989. Colonization of artificial seagrass by fishes and de- capod crustaceans: importance of proximity to natural eelgrass. J. Exp. Mar Biol. Ecol. 133:15-37. 1992. Variability in growth rates of juvenile fishes in dif- ferent estuarme habitats. Mar Ecol. Prog. Ser 85:35-53. Sogard, S. M., and K. W. Able. 1991. A comparison of eelgrass, sea lettuce macroalgae, and marsh creeks as habitat for epibenthic fishes and deca- pods. Estuarine Coastal Shelf Sci. 33:501-519. Sogard, S. M., and B. L. Gila. 1993. The infiuence of predator presence on utilization of artificial seagrass habitats by juvenile walleye pollock, Theragra chalcogramma. Environ. Biol. Fishes 37:57-65. Sogard, S. M., G. V. N. Powell, and J. G. Holmquist. 1987. Epibenthic fish communities on Florida Bay banks: relations with physical parameters and seagrass cover. Mar Ecol. Prog. Ser 40:25-39. Tanda, M. 1990. Studies on burying ability in sand and selection to the grain size for hatchery-reared marbled sole and Japa- nese flounder Nippon Suisan Gakkaishi 56(10): 1543- 1548. Turner, C. H., E. E. Ebert, and R. R. Given. 1969. Man Made Reef Ecology. Calif Dep. Fish Game, Fish Bull. 146, 221 p. Underwood, A. J. 1981. Techniques of analysis of variance in experimental marine biology and ecology. Oceanogr. Mar Biol. Ann. Rev. 19:513-605. Walker, H. J. Jr., W. Watson, and A. M. Barnett. 1987. Seasonal occurrence of larval fishes in the nearshore Southern California Bight off San Onofre, California. Estuarine Coastal Shelf Sci. 25:91-109. White, W. S. 1977. Taxonomic composition, abundance, distribution and seasonality of fish eggs and larvae in Newport Bay, California. M.A. thesis, Calif State Univ., FuUerton, Fullerton, CA, 107 p. Worthington, D. G., D. J. Ferrell, S. E. McNeill, and J. D. Bell. 1992. Effects of shoot density of seagrass on fish and deca- pods: are correlation evident over larger spatial scales? Mar Biol. 112:139-146. Yoklavich, M. M., M. Stevenson, and G. M. Cailliet. 1992. Seasonal and spatial patterns of ichthyoplankton abundance in Elkhorn Slough, California. Estuarine Coastal Shelf Sci. 34:109-126. Zar, J. H. 1984. Biostatistical analysis. Prentice-Hall, Inc., Engle- fieldCliffs, NJ, 718p. 661 Abstract.— A population model incor- porating temporal and spatial detail re- vealed that the majority of eastern Georges Bank haddock, Melanogram- mus aeglefinus, were found on the Ca- nadian side of the Canada-U.S. bound- ary-. During spring they were more wide- spread across the top of the bank and subsequently migrated eastward so that by fall almost all haddock were found in the deeper waters on the Ca- nadian side. There is a return mi- gration to the top of the bank during the winter The seasonal distribution and migration of haddock has remained stable since 1985 and migration rates do not appear to be related to the ob- served range of abundance. The distri- bution pattern since 1985 appears simi- lar to that observed between 1972 and 1984. In contrast, during 1963-71 had- dock were more widespread throughout the area in both spring and fall. Abun- dance of haddock in the Georges Bank and Gulf of Maine area was exception- ally high in the earlier period, and had- dock from the spawning component in the Great South Channel area may have accounted for a greater augmen- tation to the eastern Georges Bank population. In implementing strategies for managing this transboundary re- source, scientists will need to evaluate the nature of haddock distributions in order, in turn, to evaluate the implica- tions of their strategies. Movements of haddock, Melanogrammus aeglefinus, on eastern Georges Bank determined from a population model incorporating temporal and spatial detail Lutgarde A.M. Van Eeckhaute Stratis Gavaris Edward A. Trippel Biological Station Department of Fisheries and Oceans St Andrews, New Brunswick EOG 2X0, Canada E-mail address (for L A M Van Eeckhaute) Van EeckhauleLcfmardto-mpogcca Manuscript accepted 25 June 1998. Fish. Bull. 97:661-679 ( 1999). The haddock, Melanogrammus aeglefinus. on Georges Bank have supported a commercial fishery since the early 1900s. Since 1977, with the extension of jurisdiction by coastal states, only Canada and the United States have conducted had- dock fisheries on Georges Bank. In October 1984 the International Court of Justice (ICJ) established a maritime boundary between Canada and the United States in the Gulf of Maine area. Subsequently, fishing by Canada and the United States on Georges Bank has been re- stricted to these respective jurisdic- tions (Fig. 1). The new boundary line, referred to as the ICJ line in our study, lies over an established grid of statistical unit areas that has been used to summarize land- ings data since the 1930s. These areas have been based on, among other factors, considerations of bio- logical stock structure (principally cod, Gadus morhua, and haddock), political boundaries, and practi- calities of data collection (Halliday and Pinhorn, 1990). Aggregates of unit areas are used to determine the boundaries of fisheries "manage- ment units" — geogi'aphic areas de- fined for regulatory purposes. Prior to establishment of the maritime boundary, regulation of the haddock fishery was based on a management unit encompassing all of subarea 5, although it was recog- nized that this unit included at least two major spawning concentrations ( Walford, 1938; Bigelow and Schroe- der, 1953). Historical evidence in- dicates that the most important spawning concentration was on the eastern part of Georges Bank whereas the spawning was variable among years over the Great South Channel and the southern part of the bank (Walford, 1938), Tagging studies and other biological data have shown that little mixing occurs between haddock on Georges Bank and those in surrounding areas, e.g. 4X (Fig. 1), which is situated north and east of Georges Bank (Schuck and Arnold, 1951; Grosslein, 1962; Clark et al., 1982). Tagging studies have also indicated limited east- west movement within subdivision 5Ze. which is the eastern portion of division 5Z and includes unit areas g, h, j, m, n, and o (Fig. 1), as only about 57c of returns from haddock tagged off Cape Cod were reported from areas east of 68°W and 95*^ of returns from haddock tagged on eastern Georges Bank were re- ported from areas east of 69°W (Grosslein, 1962). In 1990, Canada adopted unit areas 5Zj and 5Zm, 662 Fishery Bulletin 97(3), 1999 4X ..-'•■ 41 Cape ■ 5Zg Cod ^1 NantucKeT Shoals 5Zo Great South Chanrjel 5Zh \ 5Zj \ Q 5'Zm ,•■■■' G ;orges Bank 5Zn 70 69 68 67 66 65 Figure 1 Fisheries statistical unit areas in NAFO subdivision 5Ze (includes 5Zg,hj,m,n.o). The Ca- nadian management unit for haddock comprises unit areas 5Zj and 5Zm (shaded!. eastern Georges Bank, as a management unit. Gavaris and Van Eeckhaute ( 1990) summarized the considerations on which the management unit bound- aries are based. The ICJ line bisects this manage- ment unit and imposes additional complexity for re- source management considerations in the absence of consistent practices by the two jurisdictions. A prerequisite to investigation of harvest strategies for the transboundary haddock resource is an under- standing of haddock distribution and migration on eastern Georges Bank in relation to the ICJ line. Earlier studies have described spatial distribution of Georges Bank haddock and patterns in relation to season, topography, time, and hydrographic condi- tions (Colton, 1955; Grosslein, 1962; Overholtz, 1985, 1987). Our study focuses on the relative distribution and net migration rates of haddock within the 5Zj,m Canadian management unit in relation to the ICJ line. Our objectives were to describe the relative abundance of haddock on the Canadian and U.S. sides of eastern Georges Bank since 1963 and to es- timate rates of migration for haddock across the ICJ line since 1985 when fishery statistics first became available at a resolution sufficient to be summarized with respect to the ICJ line. These results are neces- sary for the evaluation of the effects of regulatory measures, whether unilateral or bilateral. Methods Relative abundance Results from research vessel bottom trawl surveys were used to estimate haddock abundance on the Canadian and U.S. sides of the ICJ line on eastern Georges Bank and to subsequently derive ratios of abundance on the Canadian side to total 5Zj,m abun- dance. Annual surveys have been conducted by the U.S. National Marine Fisheries Service (NMFS) dur- ing the fall since 1963 and during the spring since 1968 and by the Canadian Department of Fisheries and Oceans (DFO) during the spring since 1986. Ratios of relative abundance were calculated to 1993 for the NMFS surveys and to 1994 for the DFO sur- veys, but survey data beyond those years were used to illustrate haddock distribution patterns as the data became available. All surveys used a stratified ran- dom design but the strata boundaries differed (Fig. 2). The strata boundaries for the DFO survey were modified to incorporate the ICJ line as a border in 1987. The results from the DFO survey in 1986, which used a different strata design, were not con- sidered in our study owing to the complication that such a design would introduce and to the limited additional information that would result. Van Eeckhaute et al Movements of Me/anogrammus oeglefinus determined from a population model 663 lll-18?m >183m, Si rata Figure 2 Strata for DFO and NMFS bottom trawl surveys with depth ranges. Several NMFS strata are bisected by the Canada-U.S. boundary line. Some strata for both countries are bisected by the 5Zj and 5Zm unit area boundaries. The DFO surveys have been conducted by the RV Alfred Needier and for one year, by its sister ship, the RV Wilfred Templeman , with a Western IIA trawl. The NMFS surveys have been conducted by the RV Albatross IV and the RV Delaware II with a BMV door until 1984 and a polyvalent door subsequently. A further gear alteration occurred with the replace- ment of the standard Yankee 36 trawl by a modified Yankee 41 trawl during the spring surveys in 1973- 81 (Hayes and BuxtonM. Conversion factors to ac- count for these differences were not considered in this study because the analysis involved only within- year comparisons. In the autumn of 1985, 1986, and 1988, both NMFS vessels made tows in the 5Zjm area but there was only one instance out of the three years, 1988, which would have resulted in slightly differ- ent ratios. The boat conversion factor was therefore not applied in our study. Abundance of haddock on each side of the ICJ line was obtained by summing the estimates of abun- dance for the strata in the respective jurisdictions. This method can be applied directly to the Canadian side for the DFO survey results because the strata boundaries incorporate the ICJ line; however, the strata on the U.S. side, 5Z3 and 5Z4, are bisected by the 5Zj,m unit area lines (Fig. 2). Several strata in the NMFS surveys are bisected by the ICJ line or the 5Zj,m unit area lines, or by both (Fig. 2). We re- fer to the parts of the bisected strata as strata sec- tions. Because the tow locations were selected at ran- dom, we computed the abundance of haddock in the Hayes, D., and N. Buxton. 1992 Assessment of the Georges Bank haddock stock in 1991. Northeast Fisheries Science Center. Res. Doc. SAW 1,3/1. (Appendix to CRD-92-02.) strata sections in the same manner as was done for entire strata according to the method in Smith ( 1988). Total abundance on either side of the ICJ line was then obtained by summing the respective abundances from strata or strata sections, or from both. Abun- dance for each stratum or stratum section was ob- tained as the average catch per tow for tows within it multipied by the number of possible tow units within it, thereby weighting the average catch per stratum or stratum section by its area. This approach could be applied in most instances; however, there were cases in the NMFS surveys where no tows were made in a stratum section ( Tables 1 and 2). Missing observations were handled in three ways; by estimation using the multiplicative model, by assumption of a zero value, or by assigning the mean per tow obtained for the whole stratum. The multiplicative model (Gavaris, 1980) was the pre- ferred method but it requires a In transformation of the catch per tow from the surveys which precludes the use of "0" values in the model; therefore, for strata sections with low abundance, this model was inap- propriate. The U.S. sections of strata 17 and 18 in both spring and fall surveys, the U.S. section of stra- tum 19 in the fall, and the Canadian section of stra- tum 18 in the spring had a predominance of "0" val- ues, suggesting that it would be reasonable to as- sume a zero value for missing observations in those strata sections. Missing observations for U.S. stra- tum section 20 within 5Zj,m were assumed to be equal to the mean catch per tow obtained for the entire stratum that extends west of 5Zjm, and the age composition for the whole stratum was extrapo- lated to that portion lying within 5Zj,m. This proce- dure was first followed for the 1991 eastern Georges 664 Fishery Bulletin 97(3), 1999 Table 1 Estimated abundance (numbers in thousands) of haddock by stratum on the Canadian and U.S. side of the ICJ line in unit areas 5Zj and 5Zm (eastern Georges Bank) from fall NMFS bottom trawl surveys. When a stratum section was not sampled, as indi- cated by "( )", a multiplicative model was used to estimate abundance except for the U.S. side of strata 17, 18, and 19 which were assigned an abundance of "0" and stratum 20 where the mean for the whole stratum was used. The last two columns show the percentage of the total abundance that came from nonsampled strata or strata sections, or from both. Year Stratum Total abundance % from non sampled Can, U.S. 16 17 18 19 20 21 22 Can. U.S. Can. U.S Can. U.S. U.S. U.S. Can. U.S. Can. U.S. Can. U.S. 1963 38,652 32,353 2148 0 (12,942) 0 30,950 4313 16,462 1942 21,761) 3772 91,964 73,330 38 0 1964 31,485 45,050 667 62 358 0 45,126 8627 2095 (627) (4535) (.555) 39,140 100,048 12 1 1965 11,434 7102 914 (0) 198 0 16.341 1849 2091 113 923 126 15,560 25,.530 0 0 1966 7999 574 123 16 57 0 7034 3817 204 8 137 175 8521 11,624 0 0 1967 3.543 344 131 0 102 5 1118 154 105 (50) 137 (45) 4018 1716 0 6 1968 4831 0 97 0 119 (0) 0 34 0 8 1.56 (16) 5203 58 0 27 1969 387 168 87 0 21 (0) 325 548 21 (12) 50 0 566 1052 0 1 1970 483 47 392 19 128 (01 5005 0 242 8 125 (31) 1369 5111 0 1 1971 451 836 319 0 13 0 758 0 32 315 87 (39) 902 1948 0 2 1972 2313 3300 312 (0) 13 0 0 0 189 8 (375) 11 3202 3319 12 0 1973 11,515 230 735 0 19 0 0 0 347 (71) (511) (63) 13,128 363 4 37 1974 387 72 58 0 13 0 0 (01 284 8 0 66 742 145 0 0 197.5 6212 9756 602 (0) 64 0 108 5512 21 (92) 175 (81) 7073 15,.549 0 1 1976 58,353 143 580 0 102 0 0 (0) 11,452 (236) (1708) 55 72,195 435 2 54 1977 2641 42 421 0 (789) 0 22 0 13,809 129 1614 (162) 19,275 355 4 46 1978 6035 2374 4571 0 543 0 3809 126 821 0 271 55 12,240 6364 0 0 1979 7173 451 416 8 1465 0 22 9 491 (.362) 6595 448 16,140 1299 0 28 1980 1643 112 3727 0 192 0 433 154 453 1200 1143 55 7157 19.54 0 0 1981 3704 1865 1558 210 172 0 81 23 274 (180) 648 66 6356 2425 0 7 1982 .591 36 1228 (0) 511 0 0 17 21 (29) 6 33 2358 115 0 25 1983 1369 947 735 (0) 192 0 0 0 158 (104) (749) 77 3202 1127 23 9 1984 414 0 2133 (0) 141 0 0 0 116 (43) (311) 55 3114 98 10 44 1985 .5315 7231 588 (0) 324 0 0 0 568 (139) 25 (123) 6819 7493 0 3 1986 6845 0 892 (0) 358 0 0 0 2274 40 12 0 10,.381 40 0 0 1987 81 179 271 (01 115 0 135 103 316 0 312 0 1094 417 0 0 1988 2537 143 493 (0) 741 0 0 0 1179 (78) 362 0 .5311 222 0 35 1989 852 47 1190 (0) (311) 5 0 103 1379 16 573 16 4447 176 7 0 1990 564 915 1204 0 345 (0) 0 0 1165 (120) 170 (106) 3448 1141 0 20 1991 483 245 609 (0) (1241 0 (0) 0 0 315 274 0 1491 560 8 0 1992 709 2869 1219 50 64 (0) 0 228 147 0 75 (39) 2214 3188 0 1 1993 1519 0 667 (0) 0 (0) 0 34 8041 (35) 162 (31) 10.390 99 0 66 Bank haddock assessment (Gavaris and Van Eeck- haute, 1991 ) and therefore the method used then was followed in our study. For all other missing observa- tions that occurred in strata which did not have a pre- dominance of zero values, a multiplicative model em- ploying strata and years as factors influencing the mean catch per tow was used to derive predicted values: In /, , = In /;-,. ^ (In PJX, -h 5^ (In P, )Z, + f , , , ■s y where s = stratum; V = year; / = survey catch/tow; P - relative power for stratum or year; X = dummy variables indicating the stratum and year of observation; and e - independent identically distributed nor- mal error. Strata sections 17u, 18c, and 18u in the spring and strata sections 17u, 18u, and 19 in the fall were not included in the multiplicative model analysis because of the prevalence of zeros in these strata sections. Van Eeckhaute et al ; Movements of Melanogrammus aeglefinus determined from a population model 665 Table 2 Estimated abundance (numbers in thousands) of haddock by stratum on the Canadian and U.S. sides of the ICJ Hne in unit areas 5Zj and 5Zm (eastern Georges Bank) from spring NMFS bottom trawl surveys. When a strata section was not sampled, as indicated by "( )", a multiplicative model was used to estimate abundance except for stratum 18 and the U.S. side of stratum 17 which were assigned an abundance of "0" and stratum 20 where the mean for the whole stratum was used. The last two columns show the percentage of the total abundance which came from nonsampled strata or strata sections, or from both. Year Stratum Total abundance % from non sampled Can. U.S. 16 17 18 19 20 21 22 Can. U.S. Can. U.S. Can. U.S. U.S. U.S. Can. U.S. Can. U.S. Can. U.S. 1968 3318 1506 595 (0) 43 0 1515 274 32 (55) (200) 77 4186 3427 5 2 1969 2164 1119 445 0 0 0 397 34 11 (25) 87 (21) 2707 1596 0 3 1970 713 7221 218 0 0 0 252 (21) 21 (26) 75 (21) 1027 7540 0 1 1971 139 430 194 0 0 5 271 0 14 73 374 (19) 720 797 0 2 1972 1002 215 310 0 0 0 144 0 91 1534 0 (20) 1403 1913 0 1 1973 2462 1052 443 (0) 0 0 685 46 21 186 1322 0 4248 1968 0 0 1974 9582 0 116 0 4 0 2841 0 416 (19) (70) 0 10,189 2860 1 1 1975 4134 538 65 (0) 0 (0) 577 0 42 16 12 (17) 4254 1148 0 2 1976 3818 1052 629 (0) 0 10) 406 (154) 358 16 (128) 0 4933 1628 3 9 1977 3221 502 48 0 0 0 681 1061 13.115 16 0 5 16,384 2266 0 0 1978 5738 2905 48 8 6 0 606 0 4652 (60) 33 (50) 10,479 3629 0 3 1979 7043 6743 319 23 0 (0) 401 (28) 253 0 594 87 8209 7284 0 0 1980 6120 27,212 2.54 54 6 (0) 24,863 (1124) 3694 0 112 (114) 10,187 53,367 0 2 1981 14.387 8704 638 3 (01 (0) 6125 2659 1905 (246) 87 230 17,017 17,967 0 1 1982 5411 3945 416 233 6 0 397 251 1179 (113) 898 22 7910 4962 0 2 1983 1933 538 735 39 0 0 0 171 (1119) 16 87 109 3874 874 29 0 1984 2623 1255 305 (0) 0 0 433 17 105 8 37 33 3071 1746 0 0 1985 9133 574 252 31 38 0 108 0 1937 (50) 87 (41) 11.447 804 0 11 1986 6313 1521 29 0 0 0 433 0 1305 4 0 (18) 7647 1975 0 1 1987 5991 115 136 0 6 0 0 1872 126 111) 25 (9) 6284 2007 0 1 1988 1417 1894 15 0 0 (0) 54 0 284 0 0 0 1716 1948 0 0 1989 4168 5022 392 93 153 10) 27 0 1631 0 798 140) 7142 5182 0 1 1990 4268 9756 44 0 (0) 10) 0 0 1291 16 12 123) 5615 9795 0 0 1991 9341 22.54 48 (0) 0 (0) 0 0 74 0 0 0 9463 2254 0 0 1992 966 369 165 (01 0 (0) 65 11 379 ilO) 0 (8) 1510 464 0 4 1993 2791 841 551 (0) 0 10) 0 0 365 (30) 237 (25) 3944 896 0 6 Treatment of missing observations is summarized in Table 3. The multiplicative model requires a In transfor- mation of the catch per tow from the surveys. This necessitated some treatment of remaining "0" val- ues in the data. To reduce the influence of "0" val- ues, the strata means rather than individual tow data were used because there are frequent occurrences of "0" catch in the individual tow data. Consequently, a test for the significance of interaction between years and strata was not possible. Because a logarithmic transformation of the mean catch per tow is required, a constant less than the minimum catch was added to "0" values. To examine the impact of adding a con- stant to the zero observations, analyses using three different constants, 0.1, 0.01, and 0.001, were con- sidered. It was found that smaller constants exerted a greater influence on the model's output as seen by increases in Cook's D statistic (Cook, 1977). More- over, the frequency distribution of the residuals showed greater departure from a normal distribu- tion when 0.01 was added and particularly when 0.001 was added, displaying a marked skew toward negative values. Consequently, 0. 1 was added to zero values for multiplicative model computations. The smallest observed value greater than "0" was 0.2. For each survey, an age-length key combining all strata on eastern Georges Bank was applied to the length composition in each stratum or stratum sec- tion to derive the age structure. The age structures to which the estimated mean number per tow for missing values were applied were taken from the 666 Fishery Bulletin 97(3), 1999 Table 3 Treatment of missing observations in the NMFS spring and fall surveys to estimate abundance on the U.S. side and the Canadian side of unit areas 5Zj and 5Zm. c = Cana- dian side of stratum; u = U.S. side of stratum; M and M = used in multiplicative model; M = missing observations estimated from multiplicative model; 5Z = 5Z stratum mean used to fill in missing observations; 0 = missing observa- tions assigned a value of 0. Stratum Spring Fall 1991.75, for ages 1-8, a, was calculated by using vir- tual population analysis with a three-month time period, i.e. t-0.25, according to the following algo- rithm. The annual instantaneous natural mortality rate, M, was assumed constant and equal to 0.2. In all equations N refers to numbers offish and, if pre- ceded by a subscript letter, indicates the source of the loss so that ^N is the number dying from fishing, ^A'^ is the number dying from natural mortality, and i^ is the number lost or gained from migration. If not preceded by a subscript, A^ refers to population abundance. The method assumes an exponential decay model: ^.,a=A^.v«,.. where the fishing mortality, F^,^. for ages 1 to 7 is obtained by solving the catch equation using a New- ton-Raphson algorithm with the three-month com- mercial catch-at-age data, pN ,^- Ny.a = PN,^JF^_„+M)t Fyjn- ~IF,„+M>t adjacent stratum section of the same stratum. When both strata sections of a stratum had missing obser- vations, the age composition from an adjacent stra- tum was used. Haddock abundance at age on the Canadian and U.S. sides of the ICJ line were obtained by applying the age compositions of strata and strata sections to their respective total abundance and then summing the results within each jurisdiction. A ra- tio of relative abundance for the Canadian side was calculated by dividing the estimate on the Canadian side by the sum of the estimates on Canadian and U.S. sides. Instantaneous rates of net migration The availability of fisheries statistics, since 1985, at a resolution sufficient to be summarized with respect to the ICJ line, in conjunction with the relative abun- dance information described above, made it possible to derive estimates of net migration rates. Catch-at- age by quarter which was required for this analysis is reported in Gavaris and Van Eeckhaute ( 1997). Net migration rates were estimated by using a model that follows from one originally proposed by Beverton and Holt ( 1957 ). Terminal population abun- dance for each cohort at the beginning of 1994 for the 1986 to 1993 year classes and at age 8 for the 1968 to 1985 year classes was taken from Gavaris and Van Eeckhaute ( 1994 ). Population abundance in all of 5Zj,m at time y, expressed in units of years, e.g. The total number of fish dying due to natural mor- tality in each quarterly period, y, was obtained by applying the quarterly rate of natural mortality to the average population abundance during the period: M^yM MtN, where N, the average population abundance during the period, is defined as iV„ .v..(] N„Jl -tF,„+M>t {F,.„+M)t DFO surveys were conducted between 10 Febru- ary and 19 March, whereas NMFS spring surveys were conducted from 23 March to 26 April. The NMFS fall surveys were conducted between 3 October and 25 October. The distributions of haddock obtained from the spring and fall survey results were consid- ered to be representative of 1 April and 1 October, respectively. For each year, six-month net migration rates were calculated for the spring-summer period (1 April-30 September) and the fall-winter period (1 October-31 March). Since proportions of haddock on the Canadian side of the ICJ line, as mdicated from DFO and NMFS survey results, were available for only two points in time, spring and fall, subsequent calculations were done on a half-year basis to coincide with the survey Van Eeckhaute et al.; Movements o\ Melanogrammus aeglefinus determined from a population model 667 timing. The 5Zj,m population numbers at the begin- ning of quarters 2 and 4 as obtained from the quar- terly VPA were used. Numbers of haddock dying from natural mortality and from fishing were summed for quarters 2 and 3 to give 1 April numbers and for quarters 4 and 1 to give 1 October numbers. The pro- portions of haddock at ages 1 to 8 on the Canadian side of the ICJ line were averaged for the two spring surveys for 1987 to 1993. These were then combined with the 1985 and 1986 proportions from the NMFS spring survey and, along with the NMFS fall pro- portions, were applied to the 5Zj,m population abun- dance at the beginning of quarters 2 and 4 to obtain population abundance on the Canadian side: ^Can.y.a = ^ y,a^Can.y.a ' where R^^,, ^.^ - the proportion occurring on the Ca- nadian side. By convention, ^,,„ and E^,^^ is positive when the net direction of migration is towards the U.S. side and negative when it is towards the Canadian side. Results Ratios of relative abundance Although the ICJ line was not established until Oc- tober 1984, to provide a historical perspective of had- dock relative abundance with respect to the line, we make reference to Canadian and U.S. sides of east- ern Georges Bank for all surveys conducted since 1963. Because relative abundances of older haddock, greater than age 8, are difficult to interpret because of the small numbers caught, an aggregated age grouping of 9-(- was used. The number of fish dying from natural mortality on the Canadian side of the ICJ line was assumed proportional to the fraction of the population occupy- ing the Canadian side on average during the period: M^Can.x.a M Ny.a Ncan.y.a ^^Can.y.a + -^C/SA.v.o ' where -^fg,,,,, and N ^^jg^ ,^ = the average population abundances on the Canadian and USA sides, respectively, during the period. The net number of fish migrating from the Cana- dian side to the U.S. side of the ICJ line was obtained by subtracting the number offish caught by the Cana- dian fishery at age, F^can. v. «> and the number of fish dying fi-om natural mortality on the Canadian side fi'om the difference between population abundance on the Canadian side at the beginning of the two sLx month periods: \T - \T _ AT _ £-'*v.a "-"Can.v.o '^ Can.y+t .a+t N - N Instantaneous rates of migration were calculated in relation to the average abundance in 5Zj,m as follows: N..Al-e-'^- where the total mortality N Z^,„=-ln^^^^:i^^ Fall season The multiplicative model accounted for about half of the total variation in catch per tow ( mul- tiple R~=0.47 ). The stratum influence with an F-value of 7.245 andP<0.001 accounted for most of the varia- tion, but the year effect also accounted for some with an F-value of 3.247 and P<0.001. The frequency dis- tribution of residuals approximated a normal distri- bution. The strata coefficients from the multiplica- tive model give an indication of the relative abun- dances between strata sections, i.e. the higher the value, the higher the abundance. For those strata sections that were used in the model, the shading in Figure 3 was scaled to the magnitude of the strata coefficients. All Canadian sections of strata had higher coefficients than the corresponding U.S. sec- tions. The four sections on the U.S. side that were not used in the model were dominated by zero catch per tow. Canadian strata sections having depth zones of 56 to 183 m exhibited the highest coefficients. The model results indicate that the strata on the Cana- dian side had higher abundance than those on the U.S. side. For each stratum the abundances by stratum or stratum section are contained in Table 1. The domi- nant strata were 16c, 16u, and 17c. Abundance in 19u and 20u used to be quite high but in recent times almost no haddock have been caught there. In con- trast, 17c has shown an increase in its relative abun- dance in recent years in comparison to early catches, although that abundance since the 1960s has dropped substantially. Total fall abundance showed that more haddock were caught on the U.S. side during the 1960s and early 1970s, after which haddock became more abundant on the Canadian side. Numbers es- timated by the multiplicative model accounted for a significant amount of the abundance on the U.S. side 668 Fishery Bulletin 97(3), 1999 42" N 4rN Fall Ln strata coefficients E3 -2.0 to -1.5 ■ -1.5 to -1.0 □ -10 to 0 ■ Oto 0.5 420N 4rN 67 W 66 W 670W 66°W Figure 3 The relative abundances between strata as characterized by the In strata coefficients from the multiplicative model for the NMFS fall and spring surveys. Strata which have a circle ( ) around them, e.g. 17u , were not used in the multiplicative model and missing observations were given a value of "0." Stratum 20 was also not used in the model because the 5Z stratum means were used for missing observations. Strata depth zones are as follows: 19 and 20, 27-55 m; 16 and 21. 56-110 m; 17 and 22, 111-183 m; 18, >183 m. in several years but, in those years, the total abun- dance on the U.S. side was usually small. For the Ca- nadian side, in most instances, very little of the abun- dance was estimated from the multiplicative model. An examination of the fall ratios of relative abun- dance, (Fig. 4), revealed two distribution patterns in haddock age 1 and older. Between 1963 and 1972, haddock were highly variable in their relative abun- dance and were found throughout the bank. After 1971, the majority of haddock ages 1 and older were found on the Canadian side as indicated by the pre- dominance of ratios >0.75. The ratios of relative abun- dance indicated that age-0 haddock were very vari- able throughout 1963-93 and were often more abun- dant on the U.S. side. Friedman's nonparametric test revealed that the ratios for ages l-9^- during the period 1972-93 were similar but the ratios for age 0 differed from these. During the period 1963-71, the ratios for all ages, 0-9-t-, did not show persistent patterns (Table 4). We conclude then that the relative abundance of age-0 haddock across the ICJ line is different from that for ages 1-9-h but only during the period 1972- 93. Therefore, for further analysis of time trends, ages 1-9-h were combined (Fig. 5). From 1963 to 1971 had- dock of ages 1+ were distributed throughout 5Zj,m but were generally found in greater numbers on the U.S. side. Since 1972, however, very few haddock at these ages have been found on the U.S. side. Between 1963 and 1971, there was one anomolously high value, 1968 with a ratio of 0.99, but all other ratios were lower than any of the values after 1971. In con- trast to ages 1-I-, the ratios of relative abimdance of age-0 haddock were very variable and indicated a less patterned distribution, being distributed throughout 5Zj,m. All except one of the stronger year classes, 1972, 1975, 1978, 1983, 198.5, 1987, and 1992, with relative abundance ratios at age 0 of 0.34, 0.28, 0.47, 0.62, 0.40, 0.21, and 0.08, respectively, were more abundant on the U.S. side. Interestingly, the weak year classes that followed these ( 1973, 1976, 1979, 1984, 1986. 1988, and 1993) were found in greater abundance on the Canadian side with ratios of 0.85, 0.85, 0.64, 1.0, 1.0, 1.0, and 0.93, respectively A likely mechanism has not been determined. The distribution of age-l-f- haddock determined by using the catch-per-tow data from the NMFS fall sur- vey, pooled over 1985-95, are shown in Figure 6. As Van Eeckhaute et aL Movements of Melanogrammus aeglefinus determined from a population model 669 NMFS Fall Survey •63 '65 '67 '69 '71 '73 '75 '77 '79 oHBD ID DDBDninnini DlDll D DDnnBlDBD DBinr il "83 '85 IDI '87 '89 '91 '93 iDBnnnDB |3 O 4 00 5 ^6 8 9+ GBDBBD r DBD DBnBnBD DBBBDBE DBBDBBB Dl nn ni □ ID^ NMFS Spring Survey '69 '71 '73 1 ■ BD nni §4 DBU o 5 DBDni a. 6 DBDnD ^ 7 DBOr" 8 nnnni 9+ nnnnBnn Dl '75 '77 '7^) DDI DBni ZZDI n n □ n XI IDI IDI ■X> 'X5 'X7 '89 '91 '93 IDI ID IDI ID ID D D D DD DDBBBD DDDBBD DDBBBD DD BBD D D D NMFS Spring Surve\ 'X7 '89 B 1.0-0.75 B 0.74-0.5 □ 0.49-0.25 n 0.24-0 c T3 a. 1965 1970 1975 1980 1985 1990 Figure 4 Ratios of relative abundance for haddock on eastern Georges Bank aggregated into four categories from the DFO and NMFS surveys and (below) population abundance and recruitment at age 1 in 5Zj,m as estimated by sequential population analysis by Gavaris and Van Eeckhaute (1995). ^ noted from the analyses of ratios of relative abundance, in recent times few haddock have been found on the U.S. side during the fall. For comparison purposes the 1963-71 distribution determined by using NMFS fall survey data are also shown. It is evident that the 1963- 71 distribution pattern is markedly different from the 1985-95 patterns. The wide spread abundance obsei-ved west of the ICJ line in the 1963-71 period is no longer present in recent times and the aggregations on the northeast peak that were apparent from 1985 to 1995 were not evident in the earlier period. Spring season As with the fall results, the multi- plicative model accounted for about half of the total 670 Fishery Bulletin 97(3), 1999 Table 4 Probability estimates of Freidmans's (NMFS) and Canadian Department nonparametric test for randomness within ages from the National Marine Fisheries Service of Fisheries and Oceans (DFO) survey ratios of relative abundance for two age groupings. Age groups Years F-value Degrees of freedom Probability of test criterion NMFS fall sSurvey 0-9+ 1963-71 8.424 9 0.5 >P> 0.25 0-9+ 1972-93 25.185 9 0.005 >P> 0.001 1-9+ 1972-93 0.8 8 P > 0.999 NMFS spring survey 1-9+ 1968-93 28.292 8 0.001 >P DFO spring survey 1-9+ 1987-94 6.0.58 8 0.75>P>0.5 variation in catch per tow (multiple i?2=o.41). The stra- tum influence accounted for most of the variation with an F-value of 8.389 and P<0.001 but the year effect was not as strong for the spring survey as it was for fall (F-value of 1.647, 0.10.5; Fig. 7). The ratios were plotted as three groups in Figure 8. For ages 4-8, there was 672 Fishery Bulletin 97(3), 1999 Table 5 Estimated abundance ( numbers in thousar ds of haddock bv stratum on the Canadian and U.S. sides of the ICJ ine in unit areas 5Zj and 5Zm (eastern Georges Bank) from spring Canadian Department of Fisheries and Oceans bottom trawl surveys. Year Canadian side U.S, side Total abundance 5Z1 5Z2 5Z3 5Z4 Canadian side U.S. side 1987 3794 7162 103 40 10.956 143 1988 1398 12,444 5344 483 13,842 5827 1989 1874 9117 279 120 10,991 399 1990 4038 9903 1814 4214 13,941 6028 1991 2243 7961 3674 2429 10.204 6103 1992 4694 5540 2508 1507 10.234 4015 1993 2946 1879 37 2988 4825 3025 1994 15,424 11,091 167 180 26,515 347 more variability in the ratios from year to year than in the fall survey, but high values predomi- nated except from the 1968-71 period when the ratios were lower Ratios for age 1 were variable with no obvious tendency but most ages 2 and 3 ratios were greater than 0.5. The short length of the DFO survey series precluded investiga- tion of this feature for these data. The effect of total 5Zj,m abundance on the spring ratios of relative abundance was exam- ined for this period because the greatest amount of dispersion onto the bank occurs during spring. This was evaluated by plotting the NMFS spring survey ratios of relative abundance against the beginning of year popu- lation numbers aggregated into three age groupings: age 1, ages 2 and 3, and ages 4-8. Population numbers were obtained from the 1995 eastern Georges Bank haddock assess- ment (Gavaris and Van Eeckhaute, 1995). No relation between abundance and ratios of rela- tive abundance was apparent. The DFO time series was not investigated for this effect be- cause of its short time span. The distribution of age 4+ haddock deter- mined by using the catch-per-tow data from the DFO spring survey pooled over 1986 to 1996 and the NMFS spring survey pooled over 1968-71, 1972- 84, and 1985-95 are shown in Figure 6. During the 1972-84 and 1985-95 period ( 1986-96 for the DFO survey) there were aggregations throughout the Ca- nadian side of 5Zj,m, especially in the area of the northeast peak, whereas the U.S. side showed lower abundance. This was especially evident for the 1985- 95 period, but there were a number of occurrences of haddock along the southern flank of the bank in this period also. The spring distribution from 1968 to 1971 Age ■) 0 -0.25 0.25-0.50 (1.50-0.75 0.75- 1.00 —I — ■—■ JIM i I 'HI Figure 7 Age by age correlation of haddock ratios of relative abundance from the NMF.S spring survey. was markedly different from the recent distribution pattern. The northeast peak aggregations were not seen, and haddock seemed to be distributed uniformly throughout 5Zj,m. The distribution pattern during this period was strongly influenced by the exception- ally abundant 1963 year class. Instantaneous rates of net migration Haddock landings by the U.S. and Canadian fisher- ies have been affected by the spawning closure area Van Eeckhaute et a\ Movements of Melanogrammus aeglefmus determined from a population model 673 Figure 8 Relative abundance of haddock on the Canadian side of 5Zj,m by age group from the NMFS spring survey. Ratios were determined as abundance on Canadian side to total abundance in 5Zjm. which was first instituted in 1970 by the Interna- tional Commission for the Northwest Atlantic Fish- eries and which was retained by both countries fol- lowing extension of jurisdiction in 1977 (Halliday, 1988). The closure, which was put into effect from March to April during 1970-71, was extended to in- clude May in 1972 and at that time an exemption for hook fisheries was introduced. In 1987 the United States extended the period to include February and in 1993, it was further extended to include January through June. The Canadian bottom trawl fishery has traditionally had limited activity during Janu- ary and February except during 1991-93. Since 1985, when Canada and the United States have conducted haddock fisheries only on their respective sides of the ICJ line, the U.S. monthly catch of haddock in unit areas 5Zj,m (Fig. 9) generally have increased from January to June, but a sharp decline in land- ings occurred in July and the months following. Ca- nadian landings peaked in June and July, the first two months following the spawning closure period, and decreased gradually to November The 6-month net migration rates calculated for three age groups (age 1, ages 2 and 3 combined, and ages 4 to 8 combined) displayed a persistent seasonal pattern indicating net movement towards the U.S. side between October and April followed by a return to the Cana- dian side between April and October (Fig. 10). Age-1 haddock had the greatest net migration rate to the Canadian side from April to September but also the 674 Fishery Bulletin 97(3), 1999 20 15 10 _ 5 -86 — Yearly landings — ^ Average landings. 1985-93 Canadian landings 1 - Figure 9 Monthly haddock landings in 5Zj,m from 1985 to 1993. most variable.The net migration rate towards the U.S. side from October to March was often lowest at age 1, intermediate at ages 2 and 3, and gi-eatest at ages 4 and older. By ages 4+ net migration back and forth seemed to be balanced. This general pattern was sup- ported by the migration rates for the entire period ex- amined, but the trend was less distinct from April 1985 to October 1987 than from April 1988 to October 1993. Examination of the spring ratios of relative abundance showed very high values during 1985-87 for ages 4-8 haddock (Fig. 8l, indicating that there was not much movement across the ICJ line during that period. Figure 11, which shows the movement and remov- als of a weak and a strong year class, the 1984 and 1985, respectively, for ages 1 and older, illustrates the migratory nature of the 5Zj,m stock across the ICJ line. The migratory pattern is well illustrated; a portion of the stock moved to the U.S. side from Oc- tober to March and most of the stock moved back to the Canadian side from April to September. In spring the abundance on the Canadian side remained higher than on the U.S. side. The U.S. catch often consisted primarily offish that had moved across the ICJ line from the Canadian side to the U.S. side and is indi- cated in the figure by the arrows pointing to the left. No obvious pattern between net migi-ation rates and population abundance was found. The range of abun- dance, however, for this period was very limited. Van Eeckhaute et a\: Movements of Melanogrammus aeglefinus determined from a population model 675 Age I Ages 2,3 Ages 4-8 1985 1986 1987 1988 1989 1990 1991 1992 1993 1994 Figure 10 Net migration rates (expressed as annual rates) across the ICJ line for 6-month tmie periods for 5Zj,m haddock. These rates show that from April to September, net haddock migration was towards the east resulting in almost all haddock in 5Zj,m residing on the Canadian side in the fall. There was then a migration westward and across the ICJ line from October to March, a migration associated with spawning activities. Discussion Observations regarding haddock spatial distribution on Georges Bank during an earlier period (Colton, 1955) and for two dominant year classes more re- cently (Overholtz, 1985) have indicated that adult haddock are more abundant off the bank in deeper waters during the fall but are found on the bank in shallower waters during the spring. Colton (1955) reported that during spring, the greatest concentra- tions of larger haddock occur in water less than 110 m and presumed this was related to spawning activi- ties. Migration from shallower water in spring to deeper water in the fall appears to occur by the end of July. Colton (1955i reported that few haddock of any age were found between 110 m and 165 m dur- ing the July-August surveys undertaken in 1949 and 1950. Older haddock, age 5+, could be found in wa- ters deeper than 165 m in unit areas 5Zg,h and j. Overholtz (1985) also suggested that survey distri- butions of age-2 haddock indicated a movement by summer to deeper water in relation to the shallower depths generally occupied in spring, Colton (1955) and Overholtz (1985) reported some segregation of ages as fish moved into deeper water, i.e. older fish occupied deeper water than did younger fish, but Colton ( 1955 1 observed that in spring there appeared to be less segregation of age groups. Although we observed two distinct distribution patterns over time within 5Zj,m, our analysis indi- cates that the depth-related patterns described above were persistent. The 1963-71 distribution pattern will be discussed more fully later in this section. Patterns observed from 1972 to recent times show that adult haddock in 5Zj,m were broadly dispersed over the bank during the spring period, their distri- bution extending westward of the ICJ line especially along the southern flank of Georges Bank. The NMFS strata with the highest densities in the spring were the Canadian portions of 21 and 16 which have a depth range from 56 to 110 m. The ratios of relative abundance, especially those from the DFO survey, clearly show that the majority of age-2-(- haddock occupied waters within the Canadian jurisdiction at that time. Both NMFS and DFO spring surveys ex- hibit similar distribution patterns with similar ar- eas of concentration. During the fall, haddock distri- bution shifted to the east onto the deeper bank slopes and few haddock remained on the U.S. side of the ICJ line. The ratios of relative abundance indicate that virtually all age l-i- haddock within 5Zj,m are on the Canadian side of the ICJ line in the fall. 676 Fishery Bulletin 97(3), 1999 The net migration results suggest that the west- ward migration of age-2 and older haddock in the winter-spring period is matched by a corresponding eastward migration during the summer-fall period of roughly similar magnitude. The net result is that roughly 20% to 25% of the adult haddock move across the ICJ line during a cycle. There may be a tendency for older haddock to migrate back and forth across the boundary line in greater numbers than for younger haddock because the instantaneous rates of net migration during the winter period for ages 4-8 1.6 Age 1 A 1984 year class g E B Natural Mortality Fishing Mortality Migrating to Canadian side '' 1 Ag ^ ^ « Migrating to U.S. side '■0 ' r- J|r ll Age 3 r^ Canadian side 08 0.6 ^ Age 4 0.4 0.2 fflfln^ e 5 m A ge 6 Age 7 _ J. LxT^rBLo-i r- Apr-30Sep W 1 Oct-3 1 Mar U.S. side 1 ^ -^■■^ 1985 1 1986 1 1987 1 1988 i 198 9 1 199 0 1 1991 1 1992 B 1985 year class . Natural Mortality . -, D Fishing Mortality "1 Migrating to Canadian side I Migrating to U.S. side Figure 11 History of the (A) 1984 and (Bi 1985 haddock year classes in unit areas 5Zj,m. The first bar of each pair represents the population size at the begin- ning of each 6-month period with the numbers gained from migration, if any. The second bar indicates the numbers lost to natural mortality, fishing mortality and migration to the other side of the ICJ line. The number of haddock left over at the end of the time period is represented by the un- shaded portion of the second bar The haddock that migrated across the ICJ line are added to the population or contribute to fishing mortality on the side they are migrating to during the same time period. are often greater than for younger ages, i.e. six out of nine times (Fig. 10). No relationship was found between stock abundance and the 1985-93 spring net migration rates, or the 1968-93 spring ratios of relative abundance, when a greater range in stock abundance occurred. The variability seen in the magnitude of movement to the U.S. side in spring from year to year (Figs. 8 and 10) may reflect varia- tion in the timing of migration, variation in the mag- nitude of migration, sampling variation, or immigra- tion from another source. Improved understanding of the source of this variability and better estimates of the magnitude and extent of migration may be possible from repeated surveying during the winter-spring period. Distribution and migration show dif- ferences by age. The distribution of age-0 haddock was found to be differ- ent from that of older haddock, as has been previously described (Overholtz, 1985; Lough and Boltz, 1989; Pola- check et al., 1992). Haddock eggs and larvae spawned on the northeast peak may be distributed as far south as the Middle Atlantic Bight (Polacheck et. al., 1992). The ratios of relative abun- dance of age-0 haddock during fall are very variable and abundance often fa- vors the U.S. side. There was a pat- tern of strong year classes being more abundant on the U.S. side whereas the weak year classes that followed were more abundant on the Canadian side. Additional research is needed to deter- mine whether there is a mechanistic explanation for this observation. Our results suggest that the distribution and movement of age-1, -2 and -3 had- dock differ from those of older haddock. In spring there was some correlation of ratios of relative abundance such that age-4 and older haddock had simi- lar patterns and age-2 and age-3 had- dock were similar whereas age-1 had- dock did not correlate with any other ages. The differences likely reflect dif- ferent migration patterns of immature and mature haddock. Maturity rates increase from ages 1 to 3 and by age 4 all haddock are mature (O'Brien and Brown, 1996; Trippel et al., 1997). Haddock biomass (age 2 and older) on all of Georges Bank from 1935 to 1960 averaged 153,000 meteric tons (t). In Van Eeckhaute et a\. Movements of Melanogrammus aeglefinus determined from a population model 677 >■'■•..*♦/•?♦ •I^ tf-' 1963-71 No./tow .0.001 • 25 • 50 , #100+ Figure 12 Fall distribution uf age 1+ haddock on Georges Bank from NMFS bottom-trawl surveys. Data have been aggregated for the three time periods shown. Note that 1969 data are not included as these data were unavailable. 1965, biomass peaked at 427,000 t due to the recruit- ment of the "very strong" 1962 and the "outstand- ing" 1963 year classes that continued to dominate the population during the late 1960s and early 1970s (Clark et al., 1982). There was a large decline, by an order of magnitude, in the biomass of haddock on Georges Bank during the late 1960s and the early 1970s to a low of 14,000 t in 1973 (Clark et al., 1982 ). The fall distribution pattern during this period of exceptionally high but rapidly decreasing abundance indicates that haddock within 5Zj,m were more widely dispersed over the bank from 1963 to 1971 (Fig. 6) than that which was observed subsequently. Moreover, the ratios of relative abundance did not strongly favor the Canadian side of the ICJ line dur- ing 1963-71 (Figs. 4 and 5). This difference in distri- bution was also evident in the spring distribution pattern for ages 4 to 8 (Fig. 6) and in the spring ra- tios of relative abundance that were generally less than 0.5 during 1968-71 (Fig. 8). After this period, the ratios, though variable, were usually above 0.5. Polacheck'^ noted that haddock, since the late 1960s, were almost never caught in research survey tows in the southern and central portions of Georges Bank, areas where they had been caught regularly during the mid-1960s. The post-1971 ratio of relative abundance trends for haddock persist to the present. During the late 1970s and early 1980s, the 1975 and 1978 year classes dominated the population and since then there have been several moderately strong year classes, i.e. the 1983, 1985, 1987, and 1992 (O'Brien and Brown, 1996; Gavaris and Van Eeckhaute, 1997 ). 2 Polacheck. T. 1995. CSIRO, Division of Fisheries, P.O. Box 1538, Hobart, Tasmania 7001, Australia. Personal commun. Despite accompanjring fluctuations in haddock abun- dance, the relative abundance pattern has not re- verted to that observed from 1963 to 1971. Pola- check,- using percent zero tows and patchiness to explore spatial distribution of Georges Bank haddock, also observed that when abundance increased with the 1975 and 1978 year classes, the patterns in spa- tial distribution did not resemble those observed from 1963 to the late 1960s. The distribution pattern dur- ing that time period was influenced strongly by the exceptionally large 1963 year class and may not be typical of the stock when it was in a more stable state as during the 1935-60 period. Recognizing the two spawning components and the likelihood that had- dock may display different spatial affinities depend- ing on their origin, we hypothesize that the 1963-71 distribution patterns may have been the result of an unusually greater contribution of recruitment from the southwestern spawning component. On the other hand, since 1985, the southwestern spawning com- ponent has probably been depleted to a greater ex- tent and may not be contributing to Georges Bank production to the degree that it did from the mid- 1970s to mid-1980s. The depletion of the southwest- ern spawning component is well illustrated by the fall distribution in 5Ze of ages 1+ haddock for three time periods, 1963-71, 1972-84, and 1985-95 (Fig. 12). Note that the almost continuous distribution of haddock seen along the northern edge of the bank during the first two time periods is not apparent during the later time period. Since 1985, the relative abundance across the ICJ line and the net migration rates have likely not been strongly affected by variations in the abundance of the eastern Georges Bank spawning component. However, these attributes may get distorted by dis- 678 Fishery Bulletin 97(3), 1999 tributional overlap of the two spawning components if the abundance of the southwestern component in- creases. This overlap effect may be responsible for the higher abundance at ages 2 and 3 of the strong 1978 year class on the U.S. side in spring (Fig. 8). Note, however, that at ages 4 and older, abundance shifts to being higher on the Canadian side of 5Zj,m. From the temporal and spatial patterns of distri- bution, ratios of relative abundance with respect to the ICJ line, net migration rates, and monthly land- ings statistics, we can infer then that a seasonal migration, probably associated with spawning behav- ior, occurs in 5Zj,m. This migratory behavior may be influenced by temperature conditions but, although temperature data is collected during surveys, the time range of this information is too limited to in- vestigate the effect on migration timing. The west- ward movement of haddock onto the bank into shal- lower water probably begins around November and reaches its greatest extent to the west during the peak of spawning in March-April, after which the return migration eastwards begins. Most haddock have probably migrated eastward across the ICJ line by July and by October; haddock are found far east of the line preferring the deeper waters of the north- east slopes of the bank. An understanding and appreciation of the distri- bution and migration patterns on eastern Georges Bank in relation to the ICJ line are important pre- requisites for investigation of sustainable harvest practices for this transboundary resource. The pat- tern of migration described in our study does not al- ter sustainable levels for the area as a whole but because of the seasonal nature of exploitation by Canada and the United States, this information can provide guidance on the exploitable biomass avail- able on each side during different seasons so that consistent and equitable hai-vest rates can be deter- mined. After establishment of the international boundary, Canada and the United States have man- aged the resources in their respective territories with little attention to what was being done in adjacent waters. Though it has been suggested that sustain- able fisheries could be maintained through indepen- dent management (Gavaris et al., 1993), it is likely that the greatest potential would be achieved through consistent management. Knowledge of temporal and spatial distribution patterns can form the basis upon which consistent management practices are explored. Acknowledgments We are grateful to the staff of the National Marine Fisheries Service at the Northeast Fisheries Center, Woods Hole, Massachusetts, for their provision of research survey and commercial landings data that permitted the preparation of this manuscript. We also thank especially Ralph Halliday and Russell Brown for their valuable comments on an earlier version of the manuscript and the two anonymous referees. Literature cited Beverton, R. J. H. and S. J. Holt. 1957. On the dynamics of exploited fish populations. Fish. Inve.st. Minist. Agnc. Fish. Food UK (series 2) 19, 533 p. Bigelow, H. B., and W. C. Schroeder. 1953. Fishes of the gulf of Maine. U.S. Fish Wildl. Serv. Bull. 74, vol. 53, 577 p. Clark, S. H., W. J. Overholtz, and R. C. Hennemuth. 1982. Review and assessment of the Georges Bank and Gulf of Maine haddock fishery. J. Northw. Atl. Fish. Sci. 3: 1-27. Colton, J. B. 1955. Spring and summer distribution of haddock on Georges Bank. Special Scientific Report of the U.S. Fish and Wildlife Service — Fisheries 156, 65 p. Cook, R. D. 1977. Detection of influential observation in linear reg- ression. J. Statistical Society 74:169-174. Gavaris, S. 1980. Use of a multiplicative model to estimate catch rate and effort from commercial data. Can. J. Fish. Aquat. Sci. 37:2272-2275. Gavaris, S., and L. Van Eeckhaute. 1990. Assessment of haddock on eastern Georges Bank. Canadian Atlantic Fisheries Scientific Advisory Commit- tee (CAFSAC) Res. Doc, 90/86, 37 p. 1991. Assessment of haddock on eastern Georges Bank. CAFSAC Res. Doc. 91/86, .30 p. 1994. Assessment of haddock on eastern Georges Bank. Canadian Dep. Fisheries and Oceans (DFOl Res. Doc. 94/ 31, 38 p. 1995. Assessment of haddock on eastern Georges Bank. DFO Res. Doc. 9.5/6, 36 p. 1997. Assessment of haddock on eastern Georges Bank. DFO Res. Doc. 97/54, 72 p. Gavaris S., and L. Van Eeckhaute, M. I. Buzeta, and J. Hunt. 1993. Yield projections for the transboundary cod and had- dock resources on eastern Georges Bank. Canadian Dep. Fisheries and Oceans (DFO) Res. Doc. 93/71, 19 p. Grosslein, M. D. 1962. Haddock stocks in the ICNAF Convention Area. International Commission for the Northwest Atlantic Fish- eries (ICNAFi Redbook Part III: 124-131. Halliday, R. G. 1988. Use of seasonal spawning area closures in the man- agement of haddock fisheries in the Northwest Atlantic. Northwest Atlantic Fisheries Organization (NAFO) Sci. Coun. Studies 12:27-36. Halliday, R. G., and A. T. Pinhorn. 1990. The delimitation of fishing areas in the northwest Atlantic. J. Northwest Atl. Fish. Sci. 10:1-51. Lough, R. G., and G. R. Bolz. 1989. The movement of cod and haddock larvae onto the shoals of Georges Bank, J, Fish Biol, 35(suppl, A):71-79. Van Eeckhaute et a\ : Movements of Melanogrammus aeglefinus determined from a population model 679 O'Brien, L., and R. W. Brown. 1996. Assessment of the Georges Bank haddock stock for 1994. New England Fisheries Steering Committee (NEFSO Ref Doc. 95-13. Overholtz, W. J. 1985. Seasonal and age-specific distribution of the 1975 and 1978 year-classes of haddock on Georges Bank. Northwest Atlantic Fisheries Organization (NAFO) Sci. Coun. Stud- ies 8:77-82. 1987. Factors relating to the reproductive biology of Georges Bank haddock iMelanogrammus aeglefinus^ in 1977- 83. J. Northwest Atl. Fish, Sci. 7:145-154. Polacheck, T., D. Mountain, D. McMillan, W. Smith, and P. Berrien. 1992. Recruitment of the 1987 year class of Georges Bank haddock iMelanogrammus aeglefinus): the influence of unusual larval transport. Can. J. Fish. Aquat. Sci. 49: 484-496. Schuck, H. A., and E. L. Arnold. 1951. Comparison of haddock from Georges and Browns bank. Fish. Bull. 52:177-185. Smith, S. J. 1988. Abundance indices from research survey data. In D. Rivard (ed.), Collected papers on stock assessment meth- ods, p. 16-40. CAFSAC Res. Doc. 88/61. Trippel, E. A., M. J. Morgan, A. Frechet, C. Rollet, A. Sinclair, C. Annand, D. Beanlands, and L. Brown. 1997. Changes in age and length at sexual maturity of northwest Atlantic cod, haddock and pollock stocks, 1972- 1995. Can. Tech. Rep. Fish. Aquat. Sci. 2157:xii -fl20 p. Walford, L. A. 1938. Effects of currents on distribution and survival of the eggs and larvae of haddock iMelanogrammus aeglefinus) on Georges Bank. Bull. U. S. Bur. Fish. 49:1-73. 680 Abstract.— Growth of otoliths of adult widow rockfish, Sebastes entomelas. and yellowtail rockfish, S. flavidus. in- habiting the coastal waters off central and northern California was reduced in 1983. This reduction ( 12.69, and 20.5'^r , respectively) coincided with a strong El Nifio event that dominated oceano- graphic conditions that year. Otolith growth from 1980 to 1987 was signifi- cantly correlated with ocean tempera- t\*re, upwelling, and sea level anomaly. Specifically, in 1983, the highest ocean temperature, largest positive sea level anomaly, and lowest amount of coastal upwelling corresponded to the smallest mean otolith growth increment. Reduction of growth in otoliths of widow and yellowtail rockfish (Sebastes entomelas and 5. flavidus) during the 1983 El Nino David Woodbury Southwest Fisheries Science Center National Marine Fisheries Service, NOAA 3150 Paradise Drive Tiburon, Calilornia 94920 E-mail address David Woodburyignoaa gov Manuscript accepted 16 September 1998. Fi.sh. Bull. 97:680-689 (1999). In 1983, one of the strongest El Nino events of the century occurred in the east-central Pacific Ocean (Cane, 1983; Rasmusson, 1984; Glynn, 1988). This event had strong nega- tive effects on the growth of many marine organisms inhabiting wa- ters along the west coast of North America, including giant kelp (Zimmerman and Robertson, 1985), yellowtail rockfish (Lenarz and Wyllie Echeverria, 19861, coho and Chinook salmon (Pearcy and Schoe- ner, 1987), Pacific herring (Spratt, 1987), Pacific kelp (Germann, 1988), Pacific hake (MacLellan and Saunders, 1995), and blue rockfish (VenTresca et al., 1995). At.ypically narrow annuli (annual growth increments) formed during 1983 were observed during routine aging of otoliths from adult yellow- tail rockfish, Sebastes flavidus (Fig. 1). Because otolith growth is gener- ally proportional to fish growth, a reduction in the width of the 1983 annulus could correspond to poor fish growth caused by the El Niiio- induced anomalous oceanographic and biotic conditions present that year Widths of annually formed in- crements on the hard parts of ani- mals and plants have been used to correlate growth with a range of environmental conditions. A com- mon example is the relation be- tween the widths of tree rings and annual precipitation. In the marine environment, a reduction in annual shell growth of geoduck clams was observed to correspond with in- creased pollution (Noakes and Cambell, 1992). Boehlert et al. ( 1989) used otolith annuli to provide historical time series of growth in Sebastes that were compared with environmental factors. In a similar study, MacLellan and Saunders ( 1995 ) noted a reduction in the 1983 annulus of Pacific hake otoliths in- habiting northeastern Pacific wa- ters, attributing it to the negative effect of the 1983 El Nino. In this study, otolith growth from 1980 to 1987 was measured from two commercially important rock- fish species: yellowtail rockfish and widow rockfish (S. entomelas). These species range from British Columbia to southern California (Gunderson and Sample, 1980). They are semipelagic, associated with bottom structure located on the shelf. The measurements of otolith growth were compared with physical oceanographic variables collected during the same period. Methods Otolith growth Otoliths from female yellowtail rockfish inhabiting Cordell Bank, California (Fig. 2), were collected during research cruises conducted from 1986 to 1992. As noted earlier, during otolith age determination procedures, the 1983 annulus was Woodbury: Reduction of growth in otoliths of Sebastes entome/as and S. flavidus 681 observed to be narrower than adjacent annuli. To quantify this observation, measurements of annuh were made on these otoliths, as well as additional otoliths obtained from an ongoing port sampling program. Otoliths were collected from male and female widow and yellowtail rockfish landed at Eureka, California, and fe- male widow rockfish landed at Bodega Bay, California, during 1989 and 1990. The annulus was defined as otolith growth from the end of one summer to the end of the following summer (approximately October through September). Using reflected light on the distal surface of whole otoliths, I observed that summer growth appears as a white ring and winter growth as a dark ring (Fig. 1). An- nuli from 366 otoliths were measured by using one of the two following methods. With the first method, a camera lucida was mounted atop a dissecting microscope, which was used to trace the surface annuli from 115 female yellowtail rockfish otoliths collected at Cordell Bank. A digitizing pad was used to measure the area of each annulus from the tracings. One or two otoliths per hour could be processed with this method. The remaining 251 otoliths were ana- lyzed by using an image enhancement system comprising a personal computer interfaced with a dissecting microscope and a high resolution closed-circuit video camera. A mouse-driven crosshair was used to trace the annuli displayed on the monitor This method increased produc- tion by sixfold. The area of each annulus (mm-) was square-root-transformed prior to analysis. To compare the two methods, annuli from five otoliths measured by the first method were remeasured by using the second method. The differences in the estimated growth from these five otoliths ranged from 0.2 to 11.4%, with a mean of 2.9%, which represented a minor error compared with observed interannual growth differences. Potential problems arise when measuring all the annuli in an otolith. When these rockfish species reach sexual maturity and somatic growth slows, the reduced growth rate is re- flected in their otoliths. Therefore, it would be impossible to determine if a reduction in otolith growth was caused by poor environmental con- ditions or from energy being diverted into go- nadal development. In addition, annuli formed after the onset of maturity are deposited disproportion- ately on the proximal surface of the otolith. It is easier to measure annuli deposited during the immature phase on the distal side of the otolith. To minimize Figure 1 Otoliths from female yellowtail rockfish collected at Cordell Bank. California, showing narrow 1983 annulus (arrow I: (A) 1982, (B) 1981, and (C) 1980 year classes. these concerns, measurements were restricted to the five annuli formed prior to sexual maturity. Both male and female widow rockfish have an age of 50% maturity of five years, whereas male and female yel- lowtail rockfish are mature at six and seven years. 682 Fishery Bulletin 97(3), 1999 Table 1 Sample sizes for each combina tion of rockfish species port. sex. and yearclass. Species Port Sex Yearclass 1979 1980 1981 1982 1983 Total Yellowtail Eureka Female 2 22 10 6 11 51 Yellowtail Eureka Male 1 27 19 1 10 58 .^ Yellowtail Cordell Bank Female 16 38 47 13 1 115 Widow Eureka Female 9 9 22 11 1 52 Widow Eureka Male 12 10 21 11 4 58 Widow Bodega Bay Female 7 4 15 6 0 32 Total 47 110 134 48 27 366 respectively ( Wyllie Echeverria, 1987 ). An additional concern arises from the protracted parturition sea- son exhibited by these species. Interannual differ- ences in birthdate distributions (Woodbury and Ralston, 1991) result in interannual differences in otolith growth during the first year of life. There- fore, measurements of the first annulus were not included in the analysis because they might have biased the data. Only measurements of otolith growth for the years 1980-87, comprising ages 1-4 and otoliths from fish belonging to the 1979—83 year classes (Table 1) were used. The results of this study are dependent upon the accuracy of the ages assigned to each otolith. It is not uncommon for ages to vary by one or more years between readers or among readers re-aging the same otolith. The 1983 annulus was observed over a pe- riod of several years during which one additional annulus was deposited each subsequent year for year classes prior to 1983. By using the distinctive 1983 annulus (Fig. 1) as a natural tag (MacLellan and Saunders, 1995), I was able to assign an age to each otolith in this study that was accurate. To test the validity of the 1983 annulus in this study, twelve otoliths that had previously been aged were re-examined without knowledge of their date of capture. On the basis of only the pattern of annuli widths, 11 of the 12 otoliths were assigned to their previously determined year class. The one misinter- pretation was for an otolith from a fish bom in 1984, which did not have the 1983 annulus. This error would be unlikely in the data used for this study because the otoliths were first aged by the break-and-bum method, in which the year of birth was determined from count- ing all the annuli. The narrow 1983 annulus could then be used to ensure that the correct age was obtained. The following additive model (Weisburg, 1993) was used to calculate treatment effects on annual otolith gi-owth for each species over all years. 42- 41 40- 39- 38 37 100-fm contour Eureka ^Cape Mendocino Point Arena N T Cordell Bank' Bodega Bay 8 $^ Francisco 126 125 122 121 W 124 123 Figure 2 Sampling locations at Eureka, Bodega Bay, and Cordell Bank. ■"'.;*/ ^ + «, + /^, + //,. + 4 + f,,/,- ;/,■/• where J ^,, = index of annual otolith growth of a species; \.i = mean square-root of annual growth; a^ = year "7" fixed effect; P^ = age Y fixed effect; 7,, = port "/?" fixed effect; S/ = sex "/" fixed effect; and f^ II = normal error term. This model assumes no significant interaction among any of the four main effects. This assumption was evaluated in Figures 3 and 4, where annual least- square means are plotted against year for each spe- cies-sex-port combination by using both a combined model with no interaction term for the age-by-year Woodbury: Reduction of growth in otoliths of Sebastes entomelas and 5. flavidus 683 effect, and a model with interactions that esti- mates a distinct otolith growth effect for each age- by-year combination. These figures show that although there are statistically significant inter- actions, the signal for the year effect is not strongly effected by excluding the interaction terms from the model. The additional variance explained by the model with interactions was only 3.5% and 4.2%, for yellowtail and widow rockfish, respectively. Because inclusion of an interaction term contributes little to the explana- tory power of the model, and analyses that in- clude interaction terms are much more restric- tive in terms of data requirements, the simple, combined additive model was used, thereby al- lowing the computation of a longer time series of annual otolith growth effects (1980-87). Oceanographic databases Thirteen oceanographic variables were analyzed to determine their relation with otolith growth patterns. These included eight measures of ocean temperature, three upwelling indices, a sea level statistic, and the northeast Pacific atmospheric pressure index. All ocean temperature data were obtained from the Fleet Numerical Oceanogra- phy Center, Master Oceanographic Observations Data Set (Sharp and McClain, 1993). The data in- cluded temperature at the sea surface, at 100 m, at 200 m, and the depth of the 10°C isotherm. Measurements were summarized from two coastal areas (i.e. 36-39°N x 121-125°W and 39- 42°N X 123-127°W), which together encompass the region where the fish were collected. Annual means for 1980-87 were calculated from the monthly means of October through September, the otolith growth year as defined in this study. Annual upwelling indices (Bakun, 1975) were calculated from daily upwelling statistics. In this instance, the mean of the February to April monthly means was computed, as these three months bracket the spring transition (Strub et al., 1987) when nutrient-rich water is first up- welled to the surface, initiating the biological production cycle (Gushing, 1975). Upwelling data were summarized from three localities (i.e. 36°N, 39°N, and 42°N), which represent the area offish collections. Sea level data compiled at the San Fran- cisco tidal station were used to compute monthly anomalies, after correcting for differences in atmo- spheric pressure (University of Hawaii^. Annual 3.0 2.6 2 2 - £ 2.6 2.2 1.8 Female Bodega Bay Female Eureka Combined Agel Age 2 — ■ Age 3 2.6 - 2.2 Male Eureka Combined Age1 Age 2 — - Age 3 82 83 Year 84 Figure 3 Comparison of mean annual otolith growth indices (1982-84) among models with and without interaction terms for widow rockfish. Bars represent ±1 standard error of the mean. ' Sea Level Center, Univ. Hawaii, Honolulu, Hawaii. Unpubl. data. 1994. mean sea level anomalies were then calculated from the monthly means of October through September Lastly, the northeast Pacific atmospheric pressure index is an indicator of atmospheric forcing upon ocean waters (Beamish and Bouillon, 1993). This index is the mean monthly difference in surface at- mospheric pressure between 40°N, 120°W (Reno, 684 Fishery Bulletin 97(3), 1999 Nevada) and 50^N, 170°W (Aleutian Islands) (NOAA^). A plot of the data revealed an increase in the inten- sity of the pressure differences during the winter months. Therefore, annual means were calculated from the monthly means for November to March. 4.2 3.8 3.4 1 3.0 ■ 2.6 2.2 2.2 Female Cordell Bank 3.8 ■ 3.4 C7> i 3.0 o o 2.6 22 3,8 3.4 3.0 - 2.6 Female Eureka Male Eureka — Combined Age1 -— Age 2 - - Age 3 82 83 Year 84 Figure 4 Comparison of mean annual otolith growth indices 11982-84) among models with and without interaction terms for yellow- tail rockfish. Bars represent ±1 standard error of the mean. Owing to strong interdependence among these 13 oceanographic variables, a principal component analysis (SAS Institute Inc., 1988) was conducted to reduce the dimensionality of the information and to insure statistical independence of the data. The or- dination was conducted on the correlation ma- trix, so that each variable was weighted equally. The otolith data were then compared with the prin- cipal components of the oceanographic data to quan- tify the impact of the 1983 El Nifio event on the growth of yellowtail and widow rockfish otoliths. Results Patterns of annual otolith growth were similar in widow and yellowtail rockfish from 1980 to 1987 (Fig. 5). A 12.67r and 20.5% reduction in otolith growth occurred during 1983 for widow and yellowtail rockfish, respectively. This year of reduced otolith growth was followed by en- hanced growth, especially for yellowtail rockfish. The growth estimates calculated for 1980 and 1987 were based on samples from one year, i.e. the 1979 and 1983 year classes respectively, and these estimates should be viewed with caution. Although most of the variance in otolith growth is explained by the year effect, the other effects in the model are worth noting (Table 2). There was a significant difference in otolith growth be- tween sexes for yellowtail rockfish (i.e. otoliths from females grew faster than those from males); however, this was not the case for widow rock- fish. Growth of the otolith at age differs between species (Fig. 6), which may be the result of so- matic growth rate differences between the spe- cies. In a simple simulation, similar curves were produced by varying the von Bertalanffy growth coefficient (k). Yellowtail rockfish exhibit faster growth than do widow rockfish, thus yielding relatively lower indices of otolith growth with in- creasing age. There was a significant difference in otolith growth between the two ports for both species (i.e. otoliths from fish collected in Eureka grew faster than those from Bodega). Further analysis revealed that these differences began in 1983. Figure 7 shows results of four separate ANOVAs run on each of the four species-port com- binations. In these analyses, sex and port were excluded as effects in the model. Otolith growth was similar at the two ports prior to 1983. AI- ^ Pacific Marine Environmental Laboratory, National Oceanic and Atmospheric Administration, Seattle, Washington. 1994. Unpubl. data. Woodbury: Reduction of growth in otoliths of Sebastes entomelas and S flavidus 685 Table 2 Analysis of variance results for the growth model used in this study. Source df Sum of squares Mean square F-Value P>F Widow rockfish (r-= 0.318) Model 12 19.9617 1.6635 21.57 0.0001 Year 7 13.6558 1.9508 25.29 0.0001 Age 3 2.3318 0.7773 10.08 0.0001 Port 1 0.9860 0.9860 12.78 0.0004 Sex 1 0.1064 0.1064 1.38 0.2408 Error 555 42.8046 0.0771 Corrected total 567 62.7663 Yellowtail rockfish r'^=0.542) Model 12 96.8742 8.0728 87.09 0.0001 Year 7 77.8370 11.1196 119.96 0.0001 Age 3 3.3136 1.1045 11.92 0.0001 Port 1 3.5704 3.5704 38.52 0.0001 Sex 1 3.8560 3.8560 41.60 0.0001 Error 883 81.8491 0.0927 Corrected total 895 178.7233 i 30 - 2,5 though both species exhibited significant reduction in otolith growth in 1983, the effect was more pronounced at Bodega, consistent with a decreasing effect of El Nino with increasing latitude, as was ob- served in Pacific hake (MacLellan and Saunders, 1995). Although otolith growth recovered quickly for both species at Eu- reka and for yellowtail rockfish at Bodega, otolith growth for widow rockfish collected fi-om Bodega was relatively slow to recover. Results of the principal component analysis on the oceanographic data (Table 3) show that only the first two components are very informative (eigenvalue >1; first three principal components shown). Note that the first component explained 107c of the combined variation among all 13 oceanographic variables, whereas the sec- ond component accounted for an additional 18%. As the loadings on the first principal component indicate, temperature and sea level anomaly are closely linked. Up- welling loads higher with increasing lati- tude, which may relate to increased upwelling from south to north along that section of coast. Upwelling also demonstrates an inverse relation with tempera- ture and sea level anomaly. Finally, the atmospheric pressure statistic does not contribute substantially to the first principal component, although it is much more influential in the second component. 40 35 2,0 Yellowtail Rockfish Widow Rockfish 85 86 87 80 81 82 83 84 Year Figure 5 Mean annual otolith growth indices ( 1980-87) for each species. Bars represent ±2 standard errors of the mean. A multiple regression of otolith growth on the first two principal components showed that the second component was neither significant nor informative, and it was therefore dropped from the analysis. An ordinary regression of otolith growth on the first prin- cipal component resulted in a significant relation- ship for both widow and yellowtail rockfish (P=0.033 686 Fishery Bulletin 97(3), 1999 and 0.030, respectively). Plots of mean annual otolith growth index against the first principal component score (Fig. 8) reveal the extreme conditions that pre- vailed in 1983. In 1983, the lowest measurements of otolith growth corresponded to the highest sea tem- perature, highest positive sea level anomaly, and low- est upwelling index. As previously noted, this par- ticular El Niiio was one of the strongest on record for the east-central Pacific Ocean. Its signature was noted off the California coast by high sea tempera- ture at 100 m, being the warmest observed since records were first kept starting in 1944 (Sharp and McClain, 1993 ). A converse relationship between en- hanced otolith growth and lower sea temperature, lower sea level anomaly, and higher upwelling index is not as evident, although above normal growth did occur in 1985, which was the "coldest" year. 4.0 3.5 P 30 2.5 2.0 Yellowtail rockfish -^ widow rockfish 12 3 4 Age (yr) Figure 6 Mean otolith growth indices at age for widow and yellowtail rockfish. Bars represent ±2 standard errors of the mean. 4,5 4.0 s 3.5 % g I 3,0 o 2,5 2,0 Eureka yellowtail rockfish Cordell Bank yellowtail rockfish Eureka widow rockfish Bodega Bay widow rockfish 80 81 82 83 84 Year 85 86 87 Figure 7 Mean annual otolith growth indices (1980-87) for widow and yellow- tail rockfish from two geographic areas. Bars represent ±2 standard errors of the mean. Discussion In this paper, the reduction in growth of widow and yellowtail rockfish otoliths dur- ing 1983 was quantified by measuring an- nuli surface areas. Otolith growth is pos- tulated to represent a conservative mea- sure of somatic growth. Casselman ( 1990) determined that otoliths from several spe- cies of freshwater fish grew relatively faster than their body during periods of slow body growth. Francis et al. (1993) argue that otolith growth in Pagrus auratus can occur even if somatic growth has stopped. There- fore, the reduction observed fi-om this study may indicate an even larger reduction in somatic growth, perhaps affecting reproduc- tive capacity as well. This was observed by VenTresca et al. (1995) who noted a reduc- tion in both the somatic and gonadal indices of blue rockfish (S. mystinus) inhabiting California coastal waters in 1983. It appears that the oceanographic per- turbations caused by El Niiio had a nega- tive effect on the growth of these fish. Tem- perature has been implicated as a major contributor to influencing both otolith and somatic growth. Lombarte and Lleonart (1993) stated that environmental condi- tions, mainly temperature, regulate the quantity of material deposited during the formation of the otoliths. Dorn (1992) found that growth of Pacific whiting (Merluccius productus), which inhabit similar waters, is most affected by tem- perature during younger ages. Although physical oceanographic condi- tions and otolith growth were found to be significantly correlated, the causal rela- tionship may well be based on preferred food availability. Central California repre- sents the southernmost range of both widow and yellowtail rockfish. Pearson and Hightower (1991) reported slower Woodbury: Reduction of growth in otoliths of Sebastes entomelas and 5. flavidus 687 growth of widow rockfish in this area as compared with growth from more north- erly distributed populations. This may be related to diet because more northerly dis- tributed populations consume more fish and euphausiids, whereas populations in the south feed more on gelatinous zoo- plankters. Lorz et al. (1983) reported the diet of yellowtail rockfish in Queen Char- lotte Sound comprised both fish and eu- phausiids, whereas off Washington it com- prised mainly euphausiids. Pereyra et al. ( 1969) found that fish dominated the diet of both yellowtail and widow rockfish col- lected near Astoria Canyon, off northern Oregon. Brodeur and Pearcy ( 1984) noted that yellowtail rockfish preyed on a diverse assemblage, dominated by euphausiid, fish and squid, depending on the season. Adams (1987) reported that widow rock- fish collected off northern California fed primarily on gelatinous zooplankters. At Cordell Bank, during late fall, winter, and in years of reduced productivity (e.g. 1983), yellow- tail rockfish have been found to feed primarily on gelatinous zooplankton, which has a lower nutri- tional value than their preferred diet of euphausiids (MacFarlane'^). Owing to the lack of upwelling and reduced productivity during 1983, these fish were probably forced to feed on less nutritional prey. Al- though lacking stomach analysis data for 1983, Lenarz and Wyllie Echeverria ( 1986) found a signifi- cant reduction in visceral fat of yellowtail rockfish collected at Bodega Bay during 1983 as compared with 1980, a "normal" year. An assumption of this study is that the widow and yellowtail rockfish sampled during this study did not migrate to less productive waters during 1983 and then return. Most species within the Sebastes genus spend their adult life within a relatively small area (Stanley et al., 1994). Both widow and yellowtail rock- fish lead an epibenthic schooling existence, inhabit- ing the area above benthic structures. Several tag- ging studies have been conducted on yellowtail rock- fish to study the movement of these fish. Carlson and Haight (1972) noted homing behavior in yellowtail rockfish that were captured and transplanted to dif- ferent areas in Alaska. Pearcy ( 1992) tagged 25 yel- lowtail rockfish on Heceta Bank off central Oregon and noted little displacement during a one-month period. However, Stanley et al. ( 1994) noted displace- ment of several tagged yellowtail rockfish from Alas- kan and Canadian waters after they had been at large for several years. Although no tagging studies have been conducted on widow rockfish, it is worth noting 40 - 85 3,5 ^ 86 87 Grawtti Index 84 ~~~~^i^ 80 Yellowtail rockfish Otolitti cn 85 86 ^^^--^^__^ 82 81 «° 84 ^^^^^^^ 87 Widow rockfish 83 2 0 - 4-202468 Oceanography (first PCA score) Figure 8 Mean annual otolith growth indices compared to the first principal com- | ponent annual scores of oceanographic variables. Table 3 Principal component analysis results for the thirteen oceanographic variables ( 1980- -87). Pnnc pal component 1 2 3 Eigenvalue 9.10 2.36 0.96 Percent variance 70.0 18.1 7.4 Cumulative variance 70.0 88.1 95.5 Eigenvectors Sea surface temp at 39-42 N 0.259 0.269 -0.417 Sea surface temp at 36-39 N 0.287 0.255 -0.265 100 m temp at 39-42"N 0.313 0.119 0.276 100 m temp at 36-39=N 0.304 0.053 0.382 200 m temp at 39-42 = N 0.302 -0.135 -0.321 200 m temp at 36-39 = N 0.301 -0.165 -0.299 Depth of lO'C isotherm at 39-42''N 0.301 0.259 0.129 Depth of 10°C isotherm at 36-39°N 0.301 0.141 0.349 Sea level anomaly 0.296 -0.214 0.296 Upwelling index at 42" N -0.321 -0.135 -0.024 Upwelling index at 39°N -0.271 0.357 0.035 Upwelling index at 36-N -0.148 0.524 0.238 Atmospheric pressure index 0.109 0.498 -0.239 MacFarlane, R. B. 1994. Paradise Dr., Tiburon, CA. Tiburon Laboratory, Personal commun. NMFS. 3150 688 Fishery Bulletin 97(3), 1999 that otoliths collected from fish sampled from south- em California to British Columbia show distinct growth patterns, indicating little large-scale movement. One of the best applications of the anomalous 1983 annulus is its use as a chronological marker to im- prove the accuracy of assigned ages, thus increasing the veracity of age-structured models used in stock assessments (MacLellan and Saunders, 1995). Al- though the narrow 1983 annuli was most pronounced ctn young fish at the time of the event, the ring was easily discerned from older fish when their otoliths were broken, burnt, and viewed with reflected light. Therefore, it could be used for decades to aid in the accurate determination of ages of fish born prior to 1983 (Boehlert et al., 1989). Although the poor environmental conditions present during 1983 appeared to have detrimental effects on the growth of widow and yellowtail rockfish, the ef- fects were short-lived as noted in the increased otolith growth the following years, especially for yellowtail rockfish. As these fish are long-lived (for decades), they have the capacity to absorb occasional years of lower productivity without permanent detrimental effects on the population. Acknowledgments Above all, thanks must go to Steve Ralston, whose patience and advice have been extremely valuable. Bill Lenarz, Jim Bence, and Alec MacCall provided research direction. Brian Jarvis and Don Pearson supplied the otoliths used in this study. Don Pearson also allowed me the use of his image enhancement equipment. Kenneth Baltz supplied the oceano- graphic data. 1 am grateful to Steve Ralston, Mickey Eldridge, Pete Adams, and two anonymous peer reviews for their critiques of earlier drafts of this manuscript. Literature cited Adams, P. B. 1987. The diet of widow rockfish Sebasles entomelas in northern Cahfornia. In W. H. Lenarz and D. R. Gunderson (eds.). Widow rockfish: proceedings of a workshop, Tiburon, California, December 11-12, 1980, p. 37-41. U.S. Dep. Commer., NOAA Tech. Rep. NMFS-48. Bakun, A. 1975. Daily and weekly upwelling indices, west coast of North America, 1967-1973. U.S. Dep. Commer, NOAA Tech. Rep. NMFS SSRF-693, 114 p. Beamish, R. J., and D. R. Bouillon. 1993. Pacific salmon production trends in relation to climate. Can. J. Fish. Aquat. Sci. .50:1002-1016. Boehlert, G. W., M. M. Yoklavich, and D. B. Chelton. 1989. Time series of growth in the genus Sebastes from the northeast Pacific Ocean. Fish Bull. 87:791-806. Brodeur, R. D., and W. G. Pearcy. 1984. Food habits and dietary overlap of some shelf rock- fishes (genus Sebastes) from the northeastern Pacific Ocean. Fish. Bull. 82:269-293. Cane, M. A. 1983. Oceanographic events during El Nino. Science (Wash. D.C.) 222:1189-1210. Carlson, H. R., and R. E. Haight. 1972. Evidence for a home site and homingof adult yellow- tail rockfish, Sebastes flavidus. J. Fish. Res. Board Can. 29:1011-1014. Casselman, J. M. 1990. Growth and relative size of calcified structures of fish. Trans. Am. Fish. Soc. 119:673-688. Gushing, D. H. 1975. Marine ecology and fisheries. Cambridge Univ. Press, Cambridge, 278 p. Dorn, M. W. 1992. Detecting environmental covariates of Pacific whit- ing Merluccius product us growth using a growth-increment model. Fish. Bull. 90:260-278. Francis, M. P., M. W. Williams, A. C. Pryce, S. Pollard, and S.G. Scott. 1993. Uncoupling of otolith and somatic growth in Pagrus auratus (Spandae). Fish. Bull. 91:1.59-164. Germann, I. 1988. Effects of the 1983-El Niiio on growth and carbon and nitrogen metabolism of Pleurophycus gardneri (Phaeopyceae: Laminariales) in the northeastern Pacific. Mar. Biol. 99:445-455. Glynn, P. W. 1988. El Nino-Southern Oscillation 1982-1983: nearshore population, community, and ecosystem responses. Annual Review of Ecology and Systematics 19:309-345. Gunderson, D. R., and T. M. Sample. 1980. Distribution and abundance of rockfish off Washing- ton, Oregon, and California during 1977. Mar. Fish. Rev. 42:2-16. Lenarz, W. H., and T. Wyllie Echeverria. 1986. Comparison of visceral fat and gonadal fat volumes of yellowtail rockfish, Sebastes flavidus, during a normal year and a year of El Nino conditions. Fish. Bull. 84:743- 745. Lombarte, A., and J. Lleonart. 1993. Otolith size changes related with body growth, habi- tat depth and temperature. Environ. Biol. Fishes 37:297- 306. Lorz, H. v., W. G. Pearcy, and M. Fraidenburg. 1983. Notes on the feeding habits of the yellowtail rock- fish, Se6as(e.s/7ai'iiii;s, off Washington and in Queen Char- lotte Sound, Calif Fish Game 69:33-38. MacLellan, S. E., and M. W. Saunders. 1995. A natural tag on the otoliths of Pacific hake {Merluc- cius productus) with implications for age validation and migration. In D. H. Secor, J. M. Dean, and S. E. Campana (eds.). Recent developments in fish otolith research, p. 567- 580. The Belle W. Baruch Library in Marine Science No. 19. Univ. South Carolina Press, Columbia, SC. Noakes, D. J., and A. Cambell. 1992. L'se of geoduck clams to indicate changes in the ma- rine environment of Ladysmith Harbour, British Colum- bia. Environmetrics 3:81-97. Pearcy, W. G. 1992. Movements of acoustically-tagged yellowtail rockfish Sebastes flavidus on Heceta Bank, Oregon. Fish. Bull. 90:726-735. Woodbury: Reduction of growth in otoliths of Sebastes entome/as and 5. flavidus 689 Pearcy, W. G., and A. Schoener. 1987. Changes in the marine biota coincident with the 1982-83 El Niiio in the northeast subarctic Pacific Ocean. J. Geophys. Res. 92(0:14417-14428. Pearson, D. E., and J. E. Hightower. 1991. Spatial and temporal variability in growth of widow rockfish ^Sebastes entonielast. U.S. Dep. Commen. NOAA Tech. Memo., NOAA-TM-NMFS-SWFSC-167, 43 p. Pereyra, W. T., W. G. Pearcy, and F. E. Carvey Jr. 1969. Sebastodes fJaiidus. a shelf rockfish feeding on me- sopelagic fauna, with consideration of the ecological implications. J. Fish. Res. Board Canada 26:2211-2215. Rasmusson, E. M. 1984. El Niiio: the ocean/atmospheric connection. Oceanus 27i2):5-12. SAS Institute, Inc. 1988. SAS/STAT user's guide, release 6.03 edition. SAS Institute, Inc., Cary, NC. 1028 p. Sharp, G. D., and D. R. McClain. 1993. Fisheries, El Nino-Southern oscillation and upper- ocean temperature records: an eastern Pacific example. Oceanography 6:13-22. Spratt, J. D. 1987. Variation in the growth rate of Pacific herring from San Francisco Bay. California. Calif Fish Game 73(3): 132-138. Stanley, R. D., B. M. Leaman, L. Haldorson, and V. M. O'Connell. 1994. Movements of tagged adult yellowtail rockfish. Sebastes flavidus, off the west coast of North America, Fish. Bull. 92:655-663. Strub, P. T., J. S. Allen, A. Huyer, and R. L. Smith. 1987. Large-scale structure of the spring transition in the coastal ocean off western North America. J. Geophys. Res. 92(C2):1527-1544. VenTresca, D. A., R. H. Parrish, J. L. Houk, M. L. Gingas, S. D. Short, and N. L. Crane. 1995. El Nifio effects on the somatic and reproductive con- dition of blue rockfish, Sebastes mystinus. Calif Coop. Oceanic Fish. Invest. Report 36:167-174. Weisburg, S. 1993. Using hard-part increment data to estimate age and environmental effects. Can. J. Fish. Aquat. Sci. 50:1229- 1237. Woodbury, D., and S. Ralston. 1991. Interannual variation in growth rates and back-cal- culated birthdate distributions of pelagic juvenile rockfish {Sebastes spp. ) off the central California coast. Fish. Bull. 89:523-533. Wyllie Echeverria, T. 1987. Thirty-four species of California rockfishes: maturity and seasonality of reproduction. Fish. Bull. 85(2): 229-250. Zimmerman, R. C, and D. L. Robertson. 1985. Effects of El Nino on local hydrography and growth of the giant kelp Macrocystis pyrifera, at Santa Catalina Island, California. Limnol. Oceanogr. 30:1298-1302. 690 Abstract.— Most growth models are age-dependent only. Although their modifications can be used to consider, implicitly, the seasonal growth of ani- mals and the effects of tagging, a gen- eral framework is unavailable for ex- plicitly incorporating time and time- dependent factors (i.e. ambient tem- perature and food availability I in age- dependent growth models. In this pa- per, I derived general age- and time-de- [J&ndent growth models for animals and gave a comprehensive list of special cases for age- and time-dependent growth models of von Bertalanffy, lo- gistic, and Gompertz types. Such mod- els explicitly incorporate age. time, and their dependent factors and are useful for modeling growth at age and time (e.g. from length-at-age data), incre- mental growth at age and time incre- ments (e.g. length increments at age and time increments data from tagging studies), the effects of tagging, and the effects of many population character- istics. I also examined their data re- quirements, their independence of the start of time and adjustment of esti- mates of parameters essential for en- suing applications, and concluded that age- and time-dependent growth mod- els are useful for subsequent applica- tions, if and only if they are indepen- dent of the start of time or time-homo- geneous and if estimates of their pa- rameters are properly adjusted. A scheme for such an adjustment is pro- posed and demonstrated. Finally, I used nine special cases of these general mod- els to analyze tagging data on a centro- pomid perch (Lates calcarifer (Bloch)). Such analyses suggested that tagging is antagonistic to fish growth and leads to a shrinkage of size and that L. calcarifer exhibits a strong seasonality in growth, namely its length grows fast- est at the start of autumn, grows less until a full stop in the middle of win- ter, shrinks until the middle of spring, and resumes a positive growth for an- other cvcle. General age- and time-dependent growth models for animals Yongshun Xiao CSIRO Division of Fisheries GPO Box 1538, Hobart, Tasmania 7001, Australia Present address SARDI Aquatic Sciences Centre 2 Hamra Avenue West Beach, SA 5024, Australia E-mail address xiaoyongshunapi sa govau Manuscript accepted 11 August 1998. Fish. Bull. 97:690-701 ( 1999). Most growth models relate an animal's size to its age alone, are independent of time, and are meant to be useful at all times. Some fac- tors (e.g. ambient temperature and food availability) that are known to affect the growth of animals vary with time, however. Consequently, time has been incorporated in age- dependent growth models implic- itly, to consider seasonal (Pitcher and Macdonald, 1973: Appeldoorn, 1987; Smith and McFarlane, 1990; Pauly et al., 1992; Pauly and Ga- schlitz' ) and biphasic (Soriano et al., 1992) growth of animals, and the effects of tagging (Xiao, 1994). Xiao's (1996, equations 3.0-4.2, p. 1676- 1677) deterministic extensions of the classical von Bertalanffy ( 1938 ), logistic (Verhulst, 1838), and Gom- pertz (1825) growth models also serve these purposes. Similarly, Wang (1998) derived a set of age- and time-dependent growth models for a special case of the von Bertal- anffy (1938) growth equation and even constructed distribution-free and consistent estimating functions for estimating their parameters. Although these implicit age- and time-dependent growth models can describe a set of data better than age-dependent growth models, a general framework is unavailable for an explicit incorporation of time and time-dependent factors. However, an explicit entry of age, time, and time-dependent factors into growth models is essential for studying the effects of many char- acteristics of a population (e.g. its age composition, size composition, density, and size- or age-specific mortalities) on the growth of its in- dividuals (Moulton et al., 1992; Walker et al., 1998). Indeed, much insight can be gained by examining density-dependent growth alone. This is because density-dependent growth can be effected by 1) com- pensatory decreases in natural mor- tality, which may result from a de- crease in predation, cannibalism, competition or diseases; 2 ) compen- satory increases in fecundity when food is more readily available or fe- tal mortality decreases; and 3) com- pensatory increases in growth rate when more food induces ear- lier maturity and greater fecundity for each age class (Holden, 1973). For these studies to be feasible, equations for the sizes of individual animals at age a at time ^ in a popu- lation must be coupled with those of their numbers at age a (or size) at time t. Just as an increase in dimension can reveal new horizons, an explicit incorporation of time and time-de- pendent factors in age-dependent growth models can be of great use and promise. It also poses interest- ing philosophical and practical problems. Indeed, in general, time- dependency makes age- and time- ' Pauly. D., and G. Gaschutz 1979. A simple method for fitting oscillating length growth data, with a program for pocket cal- culators. ICES Council Meeting 1979/ G:24, 26 p. Xiao: General age- and time-dependent growth models for animals 691 dependent growth models depend on the start of time and thereby renders them useless, unless the start of time is known. Of course, the start of time (if it did start at all) is unknown ( although some may settle for the Big Bang) and remains a subject of philosophi- cal debate. It is obvious, then, that, for practical pur- poses, workable age- and time-dependent growth models must be independent of the start of time. But, under what conditions are they so? How should esti- mates of their parameters be adjusted to make these models useful for subsequent applications at all times? To answer these questions, both age and time must enter a growth equation, explicitly. In this paper, I derive general age- and time-de- pendent growth models for animals and give a com- prehensive list of special cases for age- and time-de- pendent von Bertalanffy (1938), logistic (Verhulst, 1838), and Gompertz (1825) growth models. Such models explicitly incorporate age, time, and their dependent factors and are useful for modeling growth at age and time (e.g. from length-at-age data), incre- mental growth at age and time increments (e.g. from length increments at age and time increments from tagging studies >, the effects of tagging, and the ef- fects of population characteristics. I also examine their data requirements, their independence of the start of time, and adjustment of estimates of their parameters for ensuing applications. Finally, I use nine special cases of these general models to analyze data on length in- crements at age and time increments from a tagging study of a centropomid perch (Lates calcarifer (Bloch)) in the Northern Territory, Australia. General age- and time-dependent growth models Just as a formal derivation of age-dependent growth models necessitates use of ordinary differential equa- tions, a formal derivation of age- and time-dependent growth models entails use of partial differential equa- tions. This is because both age and time must be taken into explicit account. Readers unfamiliar with first order partial differential equations may wish to skip immediately to Equations 6-6.3, 10-10.3, and 14-14.3, with little loss of comprehension. Now, let 0 ti.= max{tQ,aQ-c} 692 Fishery Bulletin 97(3), 1999 for a fixed value of cei?. Because vra,^) satisfies Equation 1, we obtain dwjt) dt K(t + c,t)f{w^.(t)). t>t^. (2) Equations 1 and 2 are too general to be solved analytically. Now, I examined three of their special cases for f(y(a.f)). For each of these special cases. Table 1 describes where to find equations corresponding to various quantities of interest, and Table 2 describes where to find equations corresponding to various special cases of the solution for y(a,t). Age- and time-dependent growth models of von Bertalanffy (1938) type I If f(yt, dt (3) (4) Of its many interpretations, y,„^,/a.t) can represent the asymptotic size of an average individual as age ap- proaches infinity. Solution of Equation 4 as an initial value problem with wjt> I ,^,, = wJt^J yields w,(t) = wAt)e ' + ii'(.s + c,s)_v„,„^(s + c,s)e ' ds. t>t^ (5) If a-o,|<^ then c<0, -c>0, then /,=ao-c=/-a+a„; if a-aQ>t, then c>0, -t Equations corresponding to various quantitirs type II), and Gompertz growth equation^ Table 1 )f interest for von Bei tal anff'y type I) (VB type I), von Bertalanffy (type II) (VB Quantity VB type I VB type II Gompertz Partial derivative of y(a.tl Derivative o(wjt) Solution for w/t) Solution (or y(a,t) Equation 3 Equation 4 Equation 5 Equation 6 Equation 7 Equation 1 1 Equation 8 Equation 12 Equation 9 Equation 13 Equation 10 Equation 14 Xiao: General age- and time-dependent growth models for animals 693 Table 2 Equations correspond (type II) (VB type II), ing to various special cases of the solution for \ and Gompertz growth equations. IciJ) for von Berta anffy (type iMVBtypel), von Bertalanffy Assumption about y^^^ Ja,t) Assumption about K(a,t) VB type I VBtype II Gompertz none none 6 10 14 y„,„,.(a.^i=constant none 6.0 10.0 14.0 .y„,^j(a,^i=constant A7a,ii=constant K{a.t) = K„+Acos'^{t-t,) 6.1 10.1 14.1 v„,^^fa,?J=constant v„,^^to,/)=constant ifa- ifa- -a„>t 6.2 6.3 10.2 10.3 14.2 14.3 yiaj)-- - \Kls+a-t,s)ds 3'max - [Jmax " V^A) + « " ^ ^() )]« '° ' O - O,, > f (6.0) If K"(s+a-^sJ=K()=constant in equation 6.0, then y(a,t)- \yn y(a,|,/ -a -i-a„)Je ° a - a„ < t a - a,, > t (6.1) which is the age- and time-dependent von Bertalanffy ( 1938) growth model, or (if o-Oq or t-tQ is interpreted as time at liberty) Fabens (1965) growth model, with parameters K^ and y,„av- Since many factors (e.g. ambient water temperature and food availability) vary seasonally, the instanta- neous rate of growth of many animals K(a,t) fluctuates seasonally. If data are available on K(a,t) as a function of these factors, their relationships can be hypothesized. In reality, however, few such data are available. Nonetheless, one can still hypothesize about a temporal trend in K(a,t) and attribute it to the combined effects of all responsible factors. For example, as a first approximation, K(a,t) is seasonal because of seasonal changes in ambient water temperature and food availability and can be approximated by a sine or cosine curve. Thus, if K{s + a-t,s) Tjr A 2;r Ao+^COS — (S-^,,) in Equation 6.0, an application of the trigonometric function-difference relation gives Viaj)- sin(a)-sin(^) = 2cos —{a + P) sin —ia-p) Vm.« - [ Vnu« -yia„,t-a+ a„ )]e AT n 2nl 1 ] " n T ° T\ ° 2 "I AT n 2ff/ 1 1 l^.t't-t„\ sin— 1(-(,. cos — (-/j-- (-(„) ' " n T " t[ ' 2 " ) a -Qq t in Equation 6.0, then (note that ?-a-^-ao-^l=0, or t-tQ=a-a()) 694 Fishery Bulletin 97(3), 1999 where K,„^,^., K,„i„,y,i,a^., and a are model parameters to be estimated or specified. Clearly, the functional form ofK(a,t) serves its purpose well. This is because /r,„„,K,„,„ indicate, respectively, positive, no, and negative effects of tagging on the growth of animals; and K„„„<0, K,„i„-0, and K„„„>0 suggest, respectively, a shrinkage, cessation of growth, and a slower growth of tagged animals immediately after tagging. Age- and time-dependent growth models of von Bertalanffy (1938) type II If f(y^a,t}): y(a,t) y(a,t) Equations 1 and 2 become, respectively, oV(a,n dyiaj) + - — — = K(a,t) via J) da dt P y(aj) V (a,t) V V max ^ ' ' (7) dw \t) ^^ w (t) = Kit + c,t}— — dt wit) .y„,ax<^ + c,nj t>t.. (8) Solution of Equation 8 (a Bernoulli's equation) as an initial value problem with wjt) I ,^,^ = wjtj yields wit)' 1 / f K{s + c,s) J -i-i/p wAty r K(s + c J v„„(s + < + I '■ e - ds 3'max(S + C,s)'' t>t (9) If a-cto<^ then c<0, -c>0, then t^.=a,^-c-=t-a+aii; \fa~ti,)>t, then c>0, -c<0, then t^.=t(,. In other words. viaj)- y(aQ,t - a + a^ yit^+a-tj,,)" + e • ds J v^.J.s + a-^s)" -i/p \^ is+a-t.s)ds I K(s + a-t,s) + I e ds f K(s + a-t,! •' y„,.„(s + a-/, - r A'(^+a-^;)^ -lip a -Oq t (10) If p=l. Equations 7-10 and their special cases are reduced to age- and time-dependent growth models of logistic (Verhulst, 1838) type. Ify,„„^.(.s+a-^,s)=v„m^-=constant in Equation 10, then yiaj) = ■ 1 yL. y(a„,t-a + a„}" j 1 1 1 ylx l^max y^t„ + a-t,t„)'' j {ki^'o- \ Kis+n-l,s) I/p a -a,, t (10.1) with parameters Kq and Vnmx- If in Equation 10.0, then K{s + a-t,s) = K„ + Acos — (s-t^) via J) ( 1 AT n 2lti \ \ -Kr,ia~a,.f sin— (a-Qfi (COS— - t-t^ — ^a-a:.} y^ax Uma-x ViO^J-a+a,)' j { 1 -1/p a - a,, < t jLx yih+a-t^,f -«■„(<-(„ l-^sm^'(-(„lcosy^[(-(,-|l/-(„)) -,-1/p (10.2) where /fo'-^mav^' ^' ^^id t^ are model parameters to be estimated or specified. If /frs+o-^s)=/C,„Q,-f/s:„,„,-A'„„Je-'s-'+<'-°o'/« if a-ao<^ and Xfs+a-^s)=/i:,„„,-('ii'„,„,-/r„„Je-'s-'&'/« VL a-a^>t in Equation 10.0, then (note that t-a-^aQ-t^-^^, or ^-t (10.3) where K„j^^, K,„i,j,y,„a^, and a are model parameters to be estimated or specified. Age- and time-dependent growth models of Gompertz (1825) type If f(y{a,t)) = lim- p^O p 1- y(a.t) ^v(a,n[log,(3'„,,(a,n)-log,(.y(a,n)]. Equations 1 and 2 become, respectively, <9y(a,n dytaj da dt = /^(a,nv(a,n[log^,(Vmax'o-'»)-log.(v(a,n)] dwAt) K(t + cJ)w^t)[\ogXy^Jt + c,t))-\ogXu.\{t))]. t > t^ (11) Notice that Equation 12 can be written as a linear ordinary differential equation for log,,(u',,(^i'*. ie. as d\ogXii\(t)) dt -K{t + c,t)\QgXwM)) + K(t + c,t)\ogXy^^it + c,t)). 696 Fishery Bulletin 97(3), 1999 Thus, to obtain equations under the assumption corresponding to Equations 7-10.3, one can either take limits of equations 7-10.3 forp— >0 or solve Equation 12 directly. I chose the latter, without resort to applying the L'Hopital's rule to log-transformed quantities to evaluate these limits. Solution of Equation 12 as an initial value problem with wjt) I ,^i^. =wjt^.) yields wAt) = wUf (.5+c.s)log^(.v^a^(s+c.s))e e t>t,. (13) l?-a-aQ0, then t,.=a„-c=t-a+ai)\ if a-a,,S/, then c>0, -c<0, then ^,.= ^ (14) If y„,Q^(s+a-<,Syl=y,„m.=constant in Equation 14, then y(a,n I Ki>*a-l.tidt ^max Y(a,,,t -a + a,, ) v(^,i +a -t,t„) a-a^^ t (14.1) which is the age- and time-dependent Gompertz (1825) growth model, with parameters Kfi and_v,„„j-. If K{s + a-t,s) = Kq -I- Acos — is-t^) in Equation 14.0, then y(a,t) = ^ni'a> t-O0'"»^l '-'«--'o-ao' AT n 2«f 1 ,1 a - a,, < t a - a,>t (14.2) Xiao: General age- and time-dependent growth models for animals 697 where iTo'^'ma-v-^- ^' ^^^ ^o ^^® model parameters to be estimated or specified. If K(s+a-t,s)=K„^^^-(K,„^,-K„„„)e-''-'+"-"a>'" if a-aQ'o if a-aQ>t in Equation 14.0, then (note that t-a+aQ-to=0, or t-t^)=a-aQ) v{a,t) = { ^max ,-'>'m.xi»-'Oi-'>"i'm.«-*'iii ■'°-°"""^ll - '''msx "-'0 I-"' A'ma^ -A'm.n i|e-"-'0 "" -II a-Oa t) for an in- dividual animal or for a group of individuals, or use both segments io-a^^Kt and a-aQ>t) for a group of in- dividuals. It is, however, more convenient to use only one segment in a single analysis. Indeed, although growth parameters can be estimated by use of either segment of any of Equations 6.1, 6.2, 6.3; 10.1, 10.2, 10.3; 14.1, 14.2, 14.3, it is easier to use the segment for a-ag^ gives identical results, but it is tortuous and requires first calculating ^(^^,+0-^^,,^ Data requirement for estimation of parameters in a growth model is a function of the generality of that model: the more general it is, the more data it gener- ally requires. Equations 6, 6.0, 10, 10.0, 14, and 14.0 generally require knowledge of two ages o„ and a, time t, and two sizes yfa^f-a-i-aj^j anAyia.t) lia-a^^Kt; or knowledge of two times ^^ and t, age a, and two sizes y(^Q+a-^^,,i and yCa,^^ if a-OQ>^ By contrast, use of Equations 6.1, 10.1, and 14.1 only requires knowledge of the difference between two ages a-Qij, and two sizesy(0|-|/-a-(-a,y and y(a,^i; or of the difference between two times t-t^^. and two sizes y(fo"'"^~'''a'' and jv'(a,^i. Equation 6.1 has been widely used to model tagging data, where a^ or t^^ is interpreted as time at release, a or / as time at re- capture, a-a^^ or t-t^ as time at liberty, y(a^-^t-a+a^J or yit^+a-t, to) as size at release, and y(a,fi as size at recapture. It has also been used extensively to model size at age data (obtained, say, by ageing animals by reading marks in their hard parts I, where o,^ or ^^ is interpreted as age at birth, a or ^ as age, y{a^ t-a+a^y) OT y(tQ+a-t,t^} as size at birth, and y(a.t> as size at age. However, it is rare to know an animal's two ages and their corresponding sizes; what are commonly measured are one age and its corresponding size. Consequently, it is common practice to fit Equation 6.1 into such size-at-age data to estimate age at birth a,) or t^^, as well as the gi'owth parameters, thereby implicitly assuming, for all animals concerned, that the size at birth yfaf^t-a+a^) or ydQ+a-tJ^) is zero and that the age at birth Oq or t^ is the same. Exactly the same argument applies to Equations 6.2, 6.3, 10.1, 10.2, 10.3, 14.1, 14.2, and 14.3. Data analysis Barramundi L. calcarifer is a protandrous fish found in estuaries and other coastal areas of the Indo-West Pacific (Griffin, 1987). Between August 1977 and June 1980, 4933 barramundi with a body total-length range of about 10-100 cm were captured by a combi- nation of lure fishing, tidal trap, seine, and gill net. They were measured to the nearest centimeter, tagged with the then commonly used, but apparently physically and physiologically damaging, Floy FT-2 dart tags for fish >35 cm and FD-67 anchor tags for fish <35 cm, and released in rivers flowing into the Van Diemen Gulf and the Gulf of Carpentaria of northern Australia (Davis and Reid, 1982). Of those tagged, 312 fish of a total length of 23-92 cm (mean=60 cm, SE=13 cm) were recaptured, but only 308 are used in the analysis below owing to incom- plete recapture information. The time at liberty ranged from zero to 932 d, with a mean of 219 d (SE=211 d), and the length increment from -2 1 to 35 cm, with a mean of 6 cm (SE=8 cm). Negative incre- ments in length are often observed in a tagging ex- periment because tagged animals can shrink in size immediately after tagging. Let Qq or ^^ denote time at release, a or t, time at recapture, a-cif^ or t-t^^, time at liberty, y{a^t-a+afy) or y(t^+a-t,tf^), the length of a fish at release, yf'a.^i, 698 Fishery Bulletin 97(3), 1999 its length at recapture, and y,,,^,, its maximum length. The segments of Equations 6.1, 10.1, 14.1; Equations 6.2 (p=l), 10.2 (p=l), 14.2 (p=l); and Equations 6.3, 10.3, 14.3, all for a-0||<^ were fitted into the tagging data, by using the nonlinear least squares method, under the assumptions that T'=365.25 d, time started (i.e. time /=0) on 1 January 1960 (see below for its significance), and errors my(a,t) follow independent normal distributions, with a mean of y(a,t) and a constant variance of o^ (Table 3). A likelihood ratio test suggests that Equation 6.1 is significantly dif- ferent from Equation 6.2 (F^ 304=48.6892, P<0.0001 ) or from Equation 6.3 iF.,.^^^^=4.l238, P=O.Oni)\ Equa- tion 10.1 (p=l) is significantly different from Equa- tion 10.2 (p=l) (i^2 304=45-3460, P<0.0001) or from Equation 10.3 (p=i) (F., ,,,,^=3.3241, P=0.0373); and Equation 14.1 is significantly different from Equa- tion 14.2 (F,,.jr,4==46.8516, P<0.0001) or from Equa- tion 14.3 (^.,".^^,4=3. 5345, P=0. 0304). Thus, Equations 6.2, 6.3; Equations 10.2 (p=l), 10.3 (p=l); and Equa- tions 14.2, 14.3, and their associated estimates of pa- rameters seem adequate for describing the tagging data. Selection between equations 6.2 and 6.3, between Equations 10.2 and 10.3, and between Equations 14.2 and 14.3 by developing more general models of K(a.t) was not successful because of a lack of data. Are equations 6.1, 6.2, 6.3; 10.1, 10.2, 10.3; 14.1, 14.2, 14.3 independent ofthe start of time? An age- and time-dependent growth model is useful, if and only if it is independent ofthe start of time or if it is time-homogeneous. The reason for this is that start of time is unknown. For Equations 6.0, 10.0 and 14.0 to be useful, J K{s + a-t,s)ds if a -Qq < / or K(s + a -f,s)ds if a-a,j>^ must be independent ofthe start of time t. Obviously, Equations 6.1, 6.2, 6.3; 10.1, 10.2, 10.3; 14.1, 14.2, and 14.3 all are independent ofthe start of time, where time / appears as time differences t- t,, or t-t^. 0 0 However, interesting differences exist among them. Equations 6.1, 6.3, 10.1, 10.3, 14.1, and 14.3 apply on any time scales, without any adjustment of esti- mates of their parameters in subsequent applications because they depend on time difference t-tf^ or age difference a-a^^ only. By contrast. Equations 6.2, 10.2 and 14.2 and estimates of their parameters must be properly adjusted for this purpose. Specifically, the estimate of parameter t^ in Equations 6.2, 10.2, and 14.2 must be correctly adjusted before their subse- quent applications. To make such an adjustment, suppose that all growth parameters are estimated from tagging data by using one segment of Equation 6.2, 10.2, or 14.2 on one time scale (regression time scale. Fig. 1), with time t, parameter / (estimated), and a reference time t^, (known). Now. Equation 6.2, 10.2, or 14.2 is to be applied in a future fish stock assessment on another time scale (application time scale. Fig. 1), with time t', parameter t ' (unknown. Table 3 Estimates and (in parentheses) standard errors of parameters by fitting Equations 6.1, 6.2, 6.3; 10.1 (p=l), 10.2l/)=li, 10.3 (p=li; and 14.1, 14.2, 14.3 to the barramundi tagging data using the least squares method under the assumptions that 7=365.25 d, time started (i.e. time ^=0) on 1 January 1960, and errors inyla.t) follow independent normal distributions, with a mean of y{a,t) and a constant variance of o^. P<0.0001; n=308; — = not applicable. Equa- tion >'„,<,.v'°.'"cm) if,, or A'„,„, (d-M A or k,„„/d-') t^or aid) df,. df,. F (T^(cm-) /?-' 6.1 113.4724(9.6232) 0.00065(0.000131 2,306,30965.9769 22.8923 0.9951 6.2 114.4452(8.6686) 0.00061 (0.000111 0.00088(0.000181 62.7197 (06.8279) 4.304,20333.2870 17.4525 0.9963 6.3 110.1152(8.5316) 0.00078(0.000171 -0.00028(0.001211 28.7594(41.6473) 4,304,15801.1643 22.4343 0.9952 10.1 94.7255(3.2775) 0.00161(0.000141 — — 2,306,30895.5263 22.9443 0.9951 10.2 96.2908(3.0613) 0.00148(0.000121 0.00217 (0.000281 63.5155(06.7582) 4,304,19947.8,597 17.7884 0.9962 10.3 94.4251 (3.1934) 0.00182(0.000201 -0,00012(0.002.50) 37.3267(63.1348) 4,304,15684.0774 22.6010 0.9952 14.1 100.5592(4.8653) 0.00114(0.000141 — — 2.306.31051.6220 22.8295 0.9951 14.2 102.1717(4.5198) 0.00104(0.000111 0,00153(0.000221 63.2525(06.7674) 4.304.20202.0707 17.5654 0.9963 14.3 99.6456(4.6446) 0.00129(0.000181 -0.00028 (0.002(J3) 31.6268(51.7115) 4,304,15784.7666 22.4575 0.9952 Xiao; General age- and time-dependent growth models for animals 699 to be calculated), and a reference time / ' (known). r Both reference times must be chosen properly, such that t^=t . on ^^.'s scale and t ^_'=t, on /^,"s scale, where t . is an arbitrarily chosen time. For example, /,=0 on t 's scale; ^'=0 on / "s scale. Projection of both t =t , on t 's scale and t/= t . on ^^."s scale onto a third time scale (projection time scale. Fig. 1) to find their time difference on the third time scale z'-r... It is this time difference that is to be used to calculate t^^'. To do so. fo=f'- let t-t^=t '-t ', or t'=tM '-t =t.+ T'-T J r (p r 0 ' 0 0 r r 0 r /' tirt^'+t=t'-t^-T^'+T^. Therefore, in Equations 6.2, 10.2, and 14.2, replacement of f-t^ with t'-t^-T/+z^., of a- Oq with a'-flg', and of f-t^^ with ^'-^q' will give the correct growth models for the future fish stock assessment on the required time scale (application scale). For the barramundi growth described by equation 6.2, t^ = 62.7197 d, /,,=0 corresponds to r,.= l Janu- ary 1960 (row 2, Table 3), /,,'=0 corresponds to r/=l January 1999, then t^'-t +t/- t^=62.7197+(l January 1999 )-( January 1, 1960)^62. 7 197 + 14245^:14307. 7 197 d. Therefore, replacement of t—t with t'- Application time scale Projection time scale Regression time scale / =/. t=l. Figure 1 Relation among regression, application, and projection time scales for adjusting estimates of parameters in Equations 6.2. 10.2, and 14.2 for subsequent applications. and of t-t, 0 14307.7197, of a-a„ with a'-a^' with t'-tQ in the model concerned will give the correct growth models for the future (when time starts on 1 January 1999) bar- ramundi stock assessment on the required time scale (application scale). In this ex- ample, the third time scale (projection time scale) is, of course, calendar time. Discussion This work presents general age- and time- dependent models for the growth of animals and a comprehensive list of their useful special cases, forming a basis for obtaining quantitative informa- tion on the growth of animals experiencing changes in age, time, and age- and time-varying factors. These models have many applications. An obvious one would be to examine both the short- and long-term effects of tagging on the growth of animals by use of Equations 6.3, 10.3, and 14.3; iiC K^^^^^^ indicate, respectively, positive, no, and negative effects of tagging on the growth of animals. Similarly, i^,„„ <0, /f,„„,=0, and if„„„>0 suggest, re- spectively, a shrinkage, cessation of growth, and a slower growth of tagged animals immediately after tagging. In the case of the L. calcarifer (Fig. 2), tag- ging seems to have been antagonistic to its growth (A'„„ ,>X, ,, ) and led to a shrinkage of its size iK <0). n\ax nun ^ nun This conclusion is tentative, however, because of the large standard error of d . 0.002 f 5 ^-^^^ / ,.--•■ — 0.001 3. /,■••'■ l/j""'^ Growth P /• • -0.001 0 100 200 300 400 Time at liberty or age increment during liberty (d) Figure 2 Growth rate K(a.t>=K^,^^-lK„^^~K^Je-"'-°""" if a^„t as a function of age difference a-a„ or time difference <-/„ in Equations 6.3 (•••), 10.3 ( — ) and 14.3 ( — ), with estimates of parameters K^^^^^, /f,„„, and u in Table 1 forL. calcarifer. Another application would be to study how age- and time-dependent factors other than age and time affect the growth of animals. For example, one can hypothesize about the functional forms of KUi.t), such as K(a,t)=a.T(t-t^J^, where T(t) is ambient tempera- ture, availability of food, or pH value; t^. (a time lag or lead), and a and [3 are all parameters to be esti- mated or specified. Such a model is ideal for analyz- ing data on the length or weight of an individual animal at age and time, which may be available, say, from aquaculture operations. It might also be useful for analyzing data from mark-recapture experiments, where ambient temperature or food availability of a tagged animal is measured continuously from the time of its tagging to the time of its recapture. In- deed, if L. calcarifer had been tagged with a "smart" tag that could record ambient temperature or food availability, analysis would have been made of their 700 Fishery Bulletin 97(3), 1999 0.004- s- 0.003 X^^^x 5 / _,... -„_ \ 0.002 '■■■., \ / ^ '• ^ ' ■'' TO 0.001 ^-^ ^^\_ ••.,\ // <3 0.000 -0.001 ^%^ 01JAN60 01APR60 01JUL60 01OCT60 01JAN61 Figure 3 Growth rate K(a.t)=Kf^+A cosy . / y y ^ X X *^^Sv ~ /x s \ ^ C ^^ ^ \^ ^ "^ 2 - 0 .^! 1 1 Sep Oct Nov Dec Jan Feb 25 -, B 1995-96 season 20 1 § 15 c ■i 10 u o \ 5 - Sep Oct Nov Dec Jan Feb Tune Figure 4 The monthly catchability rates estimated by the model. Dash lines indicate 95'?- confidence intervals de -ived by the 200 bootstrap assessments and percentile methods. son. The estimates of catchability at the end of these two seasons exhibited a similar difference. The catches at the beginning of the 1995-96 season were much higher than those in the same period of the 1994-95 season I Fig. 2 ). High catchability rates may be associated with abundant stocks. Thus, it can be concluded that high catchability is expected when recruitment is good. Ye and Mohammed: Analysis of variation in catchability of Penaeus semisulcatus in waters off Kuwait 709 30 - 25 - M=1.8 A ' \ ' \ ' \ — ^ \ |20 ~~~~--V o M=2.9 ^^ ^/ ^^ ^=S "~"— .^ "• ^^ X \ \ o , ^ .^.^^^ "^N ^ X \ "^ — 15 ^^""""""^"'•^m^^ V / /^ \. . ^^ i: M=4 0 ~-^.^ ^y / \ \ 15 ^^^ / \ ^ Catcha o ~~~— — ~- ^^^^ .''''"'^N V ^"^~,^ ^^^ ^N \ ~~ ~- ^ S -' N X 5 - S s n - U ^ T 1 ■ 1 Sep Oct Nov Dec Jan Feb Month Figure 5 The effect of natural mortality on the estimation of catchability derived from Kuwait's 1995-96 season data. Another interesting point is that there was a rise in catchabihty in December for the 1994-95 season, in January for the 1995-96 season. The second catchability rate peak was even sHghtly higher than the first one in both seasons. All these results show that catchability in the P. semisulcatus fishery var- ies within a fishing season and from year to year, but that the variation patterns in different seasons are similar, with two peaks (Fig. 4). There are many biases in the estimation of natu- ral morality ( Vetter, 1988 ). The natural mortality rate used in this analysis was the median value of esti- mates from various sources. Different values of natu- ral mortality will result in different estimates of catchability. Figure 5 shows comparison of the re- sults derived from the seasonal 1995-96 data when the minimum (M=1.8/yr), median (M=2.9/yr), and maximum (M=4.0/yr) estimates of natural mortality were used. The variation of the pattern of catchability was similar within this range of natural mortality. A lower natural morality results in higher catcha- bilities and a relatively large variation in catchability within a fishing season, demonstrating the correla- tion between natural mortality and catchability. Therefore, we suggest that the variation pattern derived by this method is reliable, although absolute values may be biased by the natural mortality used. The weight, W in Equation 5, can be assigned a high or low value depending on the reliability of the auxiliary information compared with the catch data (Hilborn and Walters, 1992). Here, we set Wto6 sim- ply because there were six points of catch and effort data each month, but the auxiliary index had only a single datum point each month. Actually, the auxil- iary information was used to constrain the predicted stock's declining track, and thus control total mor- tality within a certain level. A sensitivity analysis showed that reducing W by 50% resulted in an aver- age decrease of 3.5% in catchability, ranging from 2.1% to 4.8%; increasing W by 50% leads to an in- crease in catchability of 7.4% on average, with a range of 5.3% to 9.6% (Fig. 6). The primary concern here is the pattern of catchability variation; it is not sensitive to the weighting factor in the case of Kuwait's P. semisulcatus fishery (Fig. 6). One method to deter- mine a suitable value for W is by comparing the geo- metric fits of both the catch and abundance index in the case of unknown reliability over either set of data. Discussion An analytical framework to estimate changes in catchability of single cohort fisheries has been pro- 710 Fishery Bulletin 97(3), 1999 25 20 15 •S 10 o 5 - W=l,5Wn W=Wo Sep Oct Nov Dec Jan Month Figure 6 The effect of weighting fact, W, on the estimation of catchability for the 1995-96 season. Feb vided. The results of our study show that the catch- ability of the Kuwait shrimp fishery is not constant; there were high values at the beginning of the fish- ing season, lowest values at the end. The two-peak pattern of catchability may be associated with school- ing behavior at the time of recruitment to the fish- ery and the spawning migration of the species. There is also great variation in catchability on an interannual basis because schooling is associated with high recruitment (Mathews et al., 1994) and certain environmental conditions (Penn, 1984). The general declining trend in catchability may be attributed to schooling behavior. As fishing pro- gressed, shrimp schools were depleted or dispersed, and catchability declined (Drobisheva and Aseev, 1976; Garcia and Reste, 1981; van Zalinge, 1984; Mathews et al., 1994). Reduced activity caused by low temperature is another identified cause for the decline of catchability of Spencer Gulf prawns iPenaeus latisiilcatus) in Australia during the win- ter months (June to August) (Sluczanowski, 1984) and P. semisulcatus in Kuwait (Mathews and Al- Hossaini, 1982). Both schooling and effects of decreas- ing temperature likely contributed to the variation of catchability in the Kuwait shrimp fishery. The major spawning season for P. semisulcatus is from December to April (Drobisheva and Aseev, 1976; Siddeek et al., 1989). Many penaeids undertake a short shoreward migration before spawning (Garcia, 1988; Ye, 1984). Penaeus seryiisulcatus in the Gulf of Carpentaria migrate from offshore into shallow wa- ters in the spring before spawning (Dall et al., 1990). Drobisheva and Aseev (1976) reported that P. semisulcatus in the Arabian Gulf forms prespawning schools or aggregations in the spawning area. Al- though the migration pattern of P. semisulcatus in watters off Kuwait is still not clear, the pattern of catchability strongly indicates that this species may return to Kuwait waters before spawning season and concentrate in certain areas. Such shoreward move- ment by spawning schools resulted in the highest catchability, and a peak in December or January (Fig. 4), which rapidly decreased as fishing progressed. There is also the possibility that shrimp of non- Kuwait origin migrate to Kuwait waters. El-Musa ( 1982) analyzed mark-recapture data and concluded that the P. semisulcatus released in February 1979 at Dohat Al-Zaur, close to the border with Saudi Arabia, exhibited a strong northward movement. The timing of such a migration could be affected signifi- cantly by environmental conditions associated with a spawning migration, resulting in an apparent in- crease in catchability earlier in some years than in others. Ye and Mohammed: Analysis of variation in catchability of Penaeus semtsukatus in waters off Kuwait 71 1 The schooling behavior of P. semisulcatiis was as- sociated with years of higher catches (van Zahnge, 1984; Mathews et al., 1994). When recruitment to the fishery was good, larger schools, and thus higher catchability, could be expected. The catches at the beginning of the 1995-96 season were much higher than those of the same period in the 1994-95 season ( Fig. 2 ); catchability in 1995-96 was also much higher (Fig. 4). With intensive fishing effort, the shrimp stock was reduced to a very low level by the end of the fishing season, despite its initial abun- dance. Therefore, catchability is associated with stock abundance. Although the variation in total mortality and catch cannot be explained by variation in natural mortal- ity, violation of the assumption that natural mortal- ity remains constant within a season may still be the source for biases in the estimation of monthly catchability. The relatively larger discrepancy be- tween predicted and observed catches, the predicted being always lower than the observed, in the first month (Fig. 2) may result from a departure from the natural mortality rate that is generally used. An in- flated natural mortality for the first month will give lower predicted catches because the first month's catchability was related to the initial population number, which in turn was connected to all the other estimates of catchability in the fitting scheme used in this study. The fit cannot be improved by adjust- ing catchability for the first month. Although it may be true that natural mortality is constant during a fishing season, a possible second recruitment in early September, which was not considered in our study and is probably rather weak, may compensate natu- ral death to some extent and lead to a lower net natu- ral mortality than the value we used. This study reports apparent changes in catch- ability of Kuwait's P. semisulcatiis fishery. It should be noted that M, q, CPUE, and schooling must be interrelated to a certain extent. Schooling may bias CPUE with respect to years in which it does not oc- cur, and M estimates from any method applied in Kuwait may also be influenced if schooling occurs only in certain seasons. Changes in catchability to a fishery may be associated with both fishing power and population characteristics. A better understand- ing of the mechanisms underlying these results will improve stock assessment. The variations in catchability suggest that changes in catch-rate indi- ces may not actually reflect variations in abundance. Although other results support the conclusion of this study, functional relationships among catchability and population characteristics, fleet characteristics, and environmental conditions should be studied further. Acknowledgments This study was part of the Shrimp Fisheries Man- agement Project supported by the Kuwait Institute for Scientific Research. We would like to thank James Bishop for many constructive discussions and com- ments on the manuscript, and two anonymous refer- ees for extensive reviews. The contributions of the project staff are highly appreciated. Literature cited Abdul-Ghaffar, A. R., and A. Y. Y. Al-Ghunaim. 1994. Review of the Kuwait's shrimp fishery, their devel- opment and present status. Proceedings of the technical consultation on shrimp management in the Arabian Gulf.Al Khobar. 6-8 November 1994, Saudi Arabia, p 1-26. Arreguin-Sanchez, F. 1996. Catchability: a key parameter for fish stock assess- ment. Rev. Fish Biology Fisheries 6:221-242. Atran, S. M., and J. G. Loesch. 1995. An analysis of weekly fluctuations in catchability coefficients. Fish. Bull. 93(31:562-567. Chittelborough, R. G. 1970. Studies on recruitment of the western Australian rock lobster Panulirus longipes cygnus George: density and natural mortality of juveniles. Aust. J. Mar Freshwater Res. 21:131-48. " Csirke, J. 1989. Changes in the catchability coefBcient in the Peruvian anchoveta tEngraulis nn^e/is) fishery. In D. Pauly, P. Muck, J. Mendo, and I. Tsukayama (eds.),The Peruvian upwelling ecosystem: dynamics and interactions. ICLARM Conf Pro- ceedings 18, Manila, Philippines, p 207-219. Dall, W., B. J. Hill, P. C. Rothlisberg, and D. J. Staples. 1990. The biology of the Penaeidae. Adv. Mar Biol. 27: 1-489. Drobisheva, S. S., and Y. P. Aseev. 1976. On the life cycle of Penaeus semisulcatiis /Crustacea, Decapoda. Penaeidae/ in the Persian Gulf Zoologicheskij Zhurnal 5/s/:769-771. Efron, B., and R. Tibshirani. 1986. Bootstrap methods for standard errors, confidence intervals, and other measures of statistical accuracy. Sta- tistical Sci. l(l):54-77. El-Musa, M. 1982. Migration patterns and growth rates for Penaeus semisulcatus using mark recapture experiments in Kuwait. In C. P. Mathews (ed.i. Revised proceedings of the shrimp fisheries management workshop. Kuwait Institute for Scientific Research, Kuwait, p. 172-206. Garcia, S. 1988. Tropical penaeus prawns. /« J.A. GuUand (ed.), Fish population dynamics, 2nd ed. John Wiley, Chichester, p 219-249. Garcia, S., and L. Le Reste. 1981. Life cycles, dynamics, exploitation and management of coastal penaeid shrimp stocks. FAO Fish. Tech. Pap. 203. FAO. Rome. 215 p. GuUand, J. A. 1983. Fish stock assessment: a manual of basic methods. Wiley, New York. NY, 233 p. 712 Fishery Bulletin 97(3), 1999 1989. The optimum opening date in shrimp fishing; a sen- sitivity analysis. In Proceedings of the eighth shrimp and fin fisheries management workshop. Kuwait Bull. Mar. Sci. 10:71-79. Hannah, R. W. 1995. Variation in geographic stock area, catchability, and natural mortality of ocean shrimp t Pandal us jordani ): some new evidence for a trophic interaction with Pacific hake IMerluccius productus). Can. J. Fish. Aquat. Sci. 52:1018- 1029. Hilborn, R., and C. J. Walters. ■^ 1992. Quantitative fisheries stock assessment: choice, dy- namics and uncertainty. Chapman and Hall, New York, NY, 570 p. Hill, B. J. 1985. Effect of temperature on duration of emergence, speed of movement, and catchability of the prawn Penaeus escu- lentus. In P. C. Rothhsberg, B. J. Hill, and D. J. Staples (eds.). Proceedings of the second Australian national prawn seminar. Simpson Halligan, Brisbane, Australia, p. 77- 83. MacCall, A. D. 1976. Density dependence of catchability coefficient in the California Pacific sardine, Sardinops sagax caerulea, purse seine fishery. Calif Coop. Oceanic Fish. Invest. Rep. 18:136-148. 1990. Dynamic geography of marine fish populations. Washington Sea Grant Program, Seattle, WA, 153 p. Mathews, C. P. 1994. Fisheries management: the Kuwaiti experience. Mar Fish. Rev 56( 1 1:23-30. Mathews, C. P. and M. Al-Hossaini. 1982. Stock assessment of Kuwait's shrimp populations, growth, mortality and life cycles of Penaeus semisulcatus. Metapenaeus affinis and Parapenaeopsis stylifera, and management of Kuwait's shrimp stocks. Proceedings of the third shrimp and fin fisheries management workshop, fin fisheries session, 4-5 December, 1982, p. 112-172. Mathews, C. P., M. Al-Hossaini, A. R. Abdul Ghaffar, and M. Al-Shoushani. 1987. Assessment of short-lived stocks with special reference to Kuwait's shrimp fisheries: a contrast of results obtained from traditional and recent size-based techniques. In D. Pauly and G. R. Morgan (eds.l. Length-based methods in fish- eries research, p. 147-166. ICLARM Conf Proc. 13. Mathews, C. P., S. Kedidi, J. Al-Qader, I. Al-Qader, A. H. Radhi, N. I. Fita, and A. Al-Yahya. 1994. Peneaus semisulcatus stocks of the western Gulf: ef- fects of schooling and environmental variation on interstock modelling and management of Kuwaiti, Saudi Arabian and Bahraini stocks. Proceedings of the technical consulta- tion on shrimp management in the Arabian Gulf 6-8 No- vember 1994. Al Khobar, Saudi Arabia, p. 1-47. Mathews, C. P., and M. Samuel. 1991. Management and research strategies in Kuwait's trawl fishery ICES Sci. Symp. 193:330-340. Mohammed, H. M. A., J. M. Bishop, and X. Xu. 1996. Population characteristics of green tiger prawn, Penaeus semisulcatus. in Kuwait waters prior to the Gulf War. Hydrobiologia 337:37-47. Morgan, G. R. 1974. Aspects of the population dynamics of the western rock lobster Panulirus cygnus George. II. Seasonal changes in the catchability coefficient. Aust. J. Mar. Freshwater Res. 25:249-59. 1989. Separating environmental and fisheries effects in the recruitment of Gulf shrimp. Kuwait Bull. Mar. Sci. 10: 51-59. Murphy, G. 1977. Characteristics of clupeoids. /n J. A. Gulland (ed.). Fish population dynamics, 1st ed. John Wiley and Sons, New York, NY, p. 283-308. Penn, J. W. 1984. The behaviour and catchability of some commercially exploited penaeids and their relationship to stock and recruitment. In J. A. Gulland and B. J. Rothschild (eds.), Penaeid shrimps: their biology and management. Fishing News Books, Farnham, Surrey, p. 173-186. Ricker, W. E. 1975. Computation and interpretation of biological statis- tics offish populations. Bull. Fish. Res. Board Can. 191, 382 p. Siddeek, M. S. M., and A. R. Abdul-Ghaffar. 1989. The effects of the one month delay in opening the shrimp fishing season o{ Penaeus semisulcatus in 1986 in Kuwait waters. Kuwait Bull. Mar Sci. 10:81-96. Siddeek, M. S., J. M. Bishop, M. El-Musa, A. R. Abdul-Ghaffar, and M. S. Abdulla. 1994. Possible reasons for increased landings of Kuwait's green tiger shrimp (Penaeus semisulcatus) in the late 1980s. In L. M. Chouet al. (eds.) Proceedings of the third Asian fisheries forum, Manila, Philippines, p. 224-227. Siddeek, M. S. M., M. El-Musa, and A. R. Abdul-Ghaffar. 1989. Final report of the shrimp fisheries management project Phase V (MB-70). Kuwait Institute for Scientific Research, Kuwait, Pre. No. KISR 3156. 104 p. Sluczanowski, P. R. 1984. Modelling and optimal control: a case study based on the Spencer Gulf prawn fishery for Penaeus latisulcatus Kishinouye. J. Cons. Int. Explor. Mer 41:211-225. van Zalinge, N. P. 1984. The shrimp fisheries in the Gulf between Iran and the Arabian Peninsula. In J. A. Gulland and B, J. Rothschild (eds.), Penaeid shrimps: their biology and management. Fishing News Books, Farnham, England, p 71-83. Vetter, E. F. 1988. Estimation of natural mortality in fish stocks: a review. Fish. Bull. 86( 1 ):25-43. Wassenberg, T. J., and B. J. Hill. 1990. Moulting behaviour of the tiger prawn Penaeus esculentus (Haswelll. Aust. .J. Mar. Freshwater Res. 35:561-71. Winters, G. H., and J. P. Wheeler. 1985. Interaction between stock area, stock abundance, and catchability coefficient. Can. J. Fish. Aquat. Sci. 42:989- 998. Xu, X., J. M. Bishop, H. M. A. Mohammed, and A. H. Alsaffar. 1995. Estimation of the natural mortality rate of green ti- ger prawns Penaeus semisulcatus (De Hann, 1884) in Ku- wait waters using relative abundance data. J. Shellfish Res. 14(1):179-184. Ye, C. C. 1984. The prawn iPenaeus orientalis Kishinouye) in Pohai Sea and their fishery. In J. A. Gulland and B.J. Rothschild (eds.), Penaeid shrimps: their biology and manage- ment. Fishing News Books, Farnham, England, p 49-59. Ye, Y., H. M. A. Mohammed, and J. M. Bishop. 1996. An overview of the shrimp resources and fisheries in Kuwait waters. Tech. Rep. Kuwait Institute for Scientific Research, 33 p. 713 Straying of adult sockeye salmon^ Oncorhynchus nerka, entering a non-natal hatchery Jason N. Griffith Andrew P. Hendry Thomas P. Quinn School of Fisheries University of Washington Box 357980, Seattle, Washington 98195 E-mail address (for T P Quinn, contact author) tquinnin'tisti wastiinqton edu) Salmonid fishes tend to return to their natal site ("home") for repro- duction (topic reviewed by Quinn and Dittman, 1992; Quinn, 1993). Odors learned by juveniles during freshwater residence and down- stream migration guide adults dur- ing the final stages of homing mi- gration (Hasler and Scholz, 1983: Dittman and Quinn, 1996), al- though adults also respond to spawning site characteristics ( Blair and Quinn, 1991) and odors of con- specifics (Newcombe and Hartman, 1973; Honda, 1982; Groot et al., 1986). However, some salmonids do not return to their natal site, but instead "stray" and spawn else- where (Quinn, 1993). The terms "homing" and "stray- ing" are defined by the endpoints of migration (i.e. the natal or non- natal site, respectively), but during their migration some salmonids as- cend one stream, only to later leave and spawn elsewhere (Ricker and Robertson, 1935; Ricker, 1972). Mature salmonids respond to the stimulus of imprinted odors with positive rheotaxis and move down- stream when they no longer detect home odors (Johnson and Hasler, 1980); therefore movement up a non-natal stream does not neces- sarily mean that the fish will spawn there. It may be natural for salmon migrating up complex river sys- tems to ascend non-natal streams for a brief distance before the ab- sence of homestream odors triggers negative rheotaxis. However, the behavior of some fish is not ad- equately explained by such "prov- ing" behavior. For example, 16*^^^ of the sockeye salmon {Oncorhynchus nerka ) radio-tagged by Burger et al. (1995) in Tustumena Lake that eventually spawned in a particular tributary, initially entered a differ- ent tributary and stayed there for up to one week. In addition, 21^^^ of the fish that spawned on the shore- line of the lake had previously en- tered a tributary stream. These fish might be "exploring" (actively seek- ing different sites and comparing their attributes) or "wandering" ( searching in the absence of stimuli ). Unfortunately, the principal methods for studying homing fail to distinguish straying from explor- ing or other related behavior pat- terns. One approach is to collect adult salmon fi-om spawning grounds, dis- place them to another spawning site or to a nonspawning area, and monitor their subsequent move- ments. Most of the fish returned to the site of their capture (e.g. Hart- man and Raleigh, 1964; McCart, 1970; Varnavskiy and Varnavskiy, 1985; Blair and Quinn, 1991), and the authors assumed but did not verify that the capture site was home. Another approach is to cap- ture and track adults as thev move upstream to their spawning site (e.g. Berman and Quinn, 1991; Burger et al., 1995). In these stud- ies, the final spawning location is assumed to be the natal one but this also is not verified. The primary alternative ap- proach to capturing and marking adult salmon is to mark them as juveniles and monitor the locations where they subsequently spawn (e.g. Quinn and Fresh, 1984; Quinn et al., 1991; Pascual and Quinn, 1994; Vander Haegen and Doty, 1995). The origin of these salmon is known, but once they enter a hatchery, they cannot leave and are only identified as strays after they are killed. Much of what we know about the frequency of straying is based on data from hatchery popu- lations, but these data reveal little about the processes of homing and straying and may not represent wild populations (Quinn, 1993). Information on the extent to which salmon that enter a non-natal hatchery would leave, if given the opportunity, would provide insights into migratory behavior and the potential biases of estimating the extent of straying from hatchery populations. The University of Washington hatchery (UWH) pi'ovides an excel- lent opportunity to study homing, exploring, and straying of salmon. The UWH releases chinook (O. tshawytscha ) and coho (O. kisutch) salmon smolts each spring. Sock- eye salmon are abundant elsewhere in the Lake Washington watershed, and a few enter the UWH although they are not reared there. In our study, all adult sockeye that en- tered the UWH were tagged and released just outside the hatchery. If they were proving or exploring, we would not expect them to re-en- ter the UWH. However, if they re- turned repeatedly after release, it would indicate that they would Manuscript accepted 15 September 1998. Fish. Bull. 97:713-716(1999). 714 Fishery Bulletin 97(3), 1999 probably spawn at this non-natal site. We also tagged and released chinook salmon to compare the behav- ior of salmon that originated elsewhere (the sock- eye) with that of salmon that presumably originated from the hatchery (the chinook). Methods The UWH is located approximately 10 km from Puget Sound, Washington, on the north side of Portage Bay, in the Lake Washington system (Fig. 1). It produces chinook and coho but not sockeye salmon. However, an average of eight sockeye salmon (range 4-18) en- tered the pond each fall during the years of this study (1992-97). Over this period, the Lake Washington system has had a mean total sockeye escapement of 175,893 (range 122,415-400,000).^ The main spawn- ing areas within the watershed are the Cedar River, Issaquah Creek, and Bear Creek, and all sockeye salmon spawn in the watershed upstream from the UWH outfall (Fig. 1; Hendry et al., 1996), principally in October and November. During the years of our study, an average of 1299 chinook salmon (range 458-2229) entered the UWH, principally from early October through mid-Novem- ber. Analysis of coded wire tagging (CWT) data has revealed a very high fidelity of UW salmon for the 1 Egan, R. 1998. Washington Department of Fish and Wild- life. 600 Capitol Way North. Olympia. WA 98501. Personal commun. hatchery, and very low levels of straying of non-na- tal chinook salmon into the UWH (»99% of the fish entering the hatchery had been released from it; Quinn and Dittman, 1992). We therefore assumed that chinook salmon entering the hatchery had been produced there but a few might have been produced elsewhere. Twice each week during the fall, the UWH pond was partially drained and all the salmon were seined into a small area. All sockeye salmon and a subsample of the chinook salmon (in 1994-96) were transferred from the seine to a live box in the pond. These fish were tagged with a T-bar tag on each side in the musculature just below the dorsal fin and quickly released within 10 m of the ladder leading into the hatchery. All tagged chinook salmon were males ( 1 year olds in 1994; 2 year olds in 1995 and 1996) and releases were spread throughout the course of the run. Returning tagged fish of both spe- cies were recaptured during the seining of the pond. Those in good condition were released once more out- side the hatchery but those with extensive fungus on their bodies were sacrificed. Fish that returned a third time were killed, regardless of their condition. Results Between 1992 and 1997, 48 sockeye salmon (28 males and 20 females) entered the UW hatchery and were tagged and released (Table 1). The percentage of these fish that returned to the hatchery varied from 0% to 37.5% among years with an overall mean of 20.87c (Table 1). Of the ten sockeye salmon that re- turned after their first displacement and that were displaced a second time, five (50%) returned again. The sex ratio of the sockeye salmon stra3dng into the hatchery (58% males) was similar to that of those returning after displacement (60% males). From 1994 to 1996, 132 male chinook salmon were tagged and released from the UWH (Table 1). The proportion of chinook salmon that returned after dis- placement did not differ among the three years ( x'-=0.0 15, 2 df P>0.99, overall mean=77% , Table 1 ). The percentage of sockeye salmon returning after displacement was much smaller than that of the chinook salmon (20.8% vs. 77.3%; X^=48.5, 1 df, P<0.005). Of the chinook salmon that returned after release, 65% returned a second time. Figure 1 Map of the Lake Washington drainage, showing locations of the University of Washington (UW) hatchery and the major tributaries used by spawning sockeye salmon. Insert indicates the drainage's location in northwest Washington. Discussion Our finding that most of the sockeye entering the UWH appeared to be proving or exploring rather than NOTE Griffith et al ; Straying of Oncorhynchus nerka into a non-natal fnatcfiery 715 Table 1 Number of adult sockeye and chinook salmon tagged and released from the University of Washington hatchery ( 1992-97) and numbers returning to the hatchery once and twice. Numbers in parentheses represent percentages of returning fish: percentages of those returning once are based on the total number released; percentages of those returning twice are based on the number returning once. Species Year Number tagged Numbers of fish returning Once Twice Sockeye 1992 4 0(0) 0(0) 1993 6 1 (16.7) 0(0) 1994 8 1 (12.5) 0(0) 1995 4 1 (25.0) 1 (100) 1996 18 4(22.2) 4(100) 1997 8 3(37.5) 0(0) Total 48 10(20.8) 5(50.0) Chinook 1994 32 25(78.1) 13(52.0) 1995 39 30(76.91 17(56.7) 1996 61 47(77.0) 36(76.5) Total 132 102(77.2) 66(64.7) straying in the true sense is helpful in interpreting the movements of salmon during the final stages of their homing migration. Most of the sockeye (38 of 48) that entered the UWH and were released did not return again. From our data we cannot determine which term (proving, exploring, or wandering) best describes their behavior. However, five out of ten of the sockeye that returned once returned a second time after displacement. We do not understand the motivation for the behavior of these fish but inter- pret their persistence as evidence that they would have spawned at the site and thus were considered strays in the true sense. Why did only 10 of 48 sockeye salmon re-enter the UWH? Were the others unable to locate the hatch- ery or did they die prior to re-entry? The proportion of chinook salmon returning to the UWH after dis- placement ( nVc ) was much higher than that of sock- eye salmon, suggesting that the UWH was not diffi- cult to find and re-enter after displacement. The pro- portion of chinook salmon returning in this study was similar to or higher than those reported in previous studies that displaced chinook about 5 km from the UWH (Whitman et al., 1982; Brannon et al., 1986, Quinn et al., 1988). A few of the tagged sockeye in the present study returned over two weeks after re- lease, implying that the sockeye had sufficient time to find and re-enter the UWH before they were too weak. The 48 sockeye salmon that initially entered the UWH were only 0.005% of the total run of sockeye salmon to the Lake Washington system during the years of this study. We do not infer that this extremely low rate of straying or exploring is representative of spawning sites within this or other lake systems. The absence of odors from conspecific juveniles and trace odors from adults emanating from the UWH might make it less attractive than rivers perennially used by sockeye (Groot et al., 1986), and the small dis- charge and lack of appropriate habitat for spawning might also deter non-native sockeye from entering. Likewise, the proportions of sockeye salmon stray- ing and exploring in our study may not be represen- tative of those for natural systems. Entry of sockeye salmon into UWH may differ in some respects from their behavior in natural systems but exploring and straying are characteristic of salmon. On the basis of our results, many of the salmon that enter non-natal hatcheries and that are classified as strays might have left if given the chance. Studies of straying based on recoveries in hatcheries may thus overestimate straying rates, at least in some cases (Quinn et al. 1991). The lack of quantified estimates of proving or exploring behav- ior makes it difficult to assess whether the high pro- portion of these patterns found in our study is typi- cal of sockeye or Pacific salmon in general. The dis- tinction between straying and exploring is important to the management of Pacific salmon because CWT- based straying estimates from hatchery populations are commonly used to model interactions between hatchery and wild populations (Grant, 1997), and this subject needs further research. Acknowledgments We thank William Hershberger, Glenn Yokoyama, Heather Roffey, Jainrong Chang, and undergradu- ate students for assisting with salmon collection and tagging, and Andrew Dittman, Jeffrey Silverstein, Fred Utter, and Carl Burger for comments on the manuscript. Floy Tag Co. donated tags and a tag- ging gun. Andrew Hendry was supported by the H. Mason Keeler Endowment and a Natural Sciences and Engineering Research Council of Canada post- graduate scholarship. Literature cited Berman, C. H., and T. P. Quinn. 1991. Behavioural thermoregulation and homing by spring chinook salmon, Oncorhynchus tshawytscha (Walbaumi, in the Yakima River. J. Fish Biol. 39:301-312. 716 Fishery Bulletin 97(3), 1999 Blair, G. R., and T. P. Quinn. 1991. Homing and spawning site selection by sockeye salmon (Oncorhynchus nerka) in Iliamna Lake, Alaska. Can. J. Zool. 69:176-181. Brannon, E. L., T. P. Quinn, R. P. Whitman, A. E. Nevissi, R. E. Nakatani, and C. D. McAuliffe. 1986. Homing of adult chinook salmon after brief exposure to whole and dispersed crude oil. Trans. Am. Fish. Soc. 115:823-827. Burger, C. V., J. E. Finn, and L. Holland-Bartels. 1995. Patterns of shoreline spawning by sockeye salmon in a glacially turbid lake: evidence for subpopulation differentiation. Trans. Am. Fish. Soc. 124:1-15. Dittman, A. H., and T. P. Quinn. 1996. Homing in Pacific salmon: mechanisms and ecologi- cal basis. J. Exp. Biol. 199:83-91. Grant, W. S. (editor). 1997. Genetic effects of straying of non-native hatchery fish into natural populations: proceedings of the workshop June 1-2, 1995, Seattle, Washington. U.S. Dep. Commer., NOAA, NMFS Tech. Memo. NMFS-NWFSC-30, 122 p. Groot, C, T. P. Quinn, and T. J. Hara. 1986. Responses of migrating adult sockeye salmon (Oncor- hynchus nerka) to population specific odors. Can. J. Zool. 64:926-932. Hartman, W. L., and R. F. Raleigh. 1964. Tributary homing of sockeye salmon at Brooks and Karluk lakes, Alaska. J. Fish. Res. Board Can. 21:485- 504. Hasler, A. D., and A. T. Scholz. 1983. Olfactory imprinting and homing in salmon. Springer- Verlag, Berlin, 134 p. Hendry, A. P., T. P. Quinn, and F. M. Utter. 1996. Genetic evidence for the persistence and divergence of native and introduced sockeye salmon (Oncorhynchus nerka) within Lake Washington, Washington. Can. J. Fish. Aquat. Sci. 53:82.3-832. Honda, H. 1982. On the female pheromones and courtship behaviour in the salmonids, Oncorhynchus masou and O. rhodurus. Bull. Jpn. Soc. Sci. Fish. 48(11:47-49. Johnsen, P. B., and A. D. Hasler. 1980. The use of chemical cues in the upstream migration of coho salmon, Oncorhynchus kisutch Walbaum. J. Fish Biol. 17:67-73. McCart, P. J. 1970. A polymorphic population of Oncorhynchus nerka at Babine Lake. B.C. involving anadromous (sockeye) and non-anadromous (kokanee) forms. Ph. D. diss., Univ. Brit- ish Columbia, Vancouver, 135 p. Newcombe, C, and G. Hartman. 1973. Some chemical signals in the spawning behavior of rainbow trout ( Salmo gairdneri I . J. Fish. Res. Board Can. 30:99.5-997. Pascual, M. A., and T. P. Quinn. 1994. Geographical patterns of straying in fall chinook, Oncor- hynchus tshauytscha (Walbaum), from Columbia River (USA) hatcheries. Aquacult. Fish. Manage. 25 (suppl. 2):17-30. Quinn, T. P. 1993. A review of homing and straying of wild and hatch- ery-produced salmon. Fish. Res. 18:29-44. Quinn, T. P., and A. H. Dittman. 1992. Fishes. In F. Papi (ed.). Animal homing, p. 145- 211. Chapman and Hall, London. Quinn, T. P., and K. Fresh. 1984. Homing and straying in chinook salmon (Oncorhyn- chus tshauytscha ) from Cowlitz River Hatchery, Washing- ton. Can. J. Fish. Aquat. Sci. 41:1078-1082. Quinn, T. P., A. F. Olson, and J. T. Konecki. 1988. Effects of anaesthesia on the chemosensory behaviour of Pacific salmon. .J. Fish Biol. 33:637-641. Quinn, T. P., R. S. Nemeth, and D. O. Mclsaac. 1991. Homing and straying patterns of fall chinook salmon in the lower Columbia River Trans. Am. Fish. Soc. 120: 150- 156. Ricker, W. E. 1972. Hereditary and environmental factors affecting cer- tain salmonid populations. In R. C. Simon and P. A. Larkin (eds.). The stock concept in Pacific salmon, p. 19- 160. H. R. MacMillian Lectures in Fisheries, Univ. Brit- ish Columbia, Vancouver, B.C. Ricker, W. E., and A. Robertson. 1935. Observations on the behaviorof adult sockeye salmon during the spawning migration. Can. Field-Nat. 49:132- 134. Vander Haegen, G. and D. Doty. 1995. Homing of coho and fall chinook salmon in Washing- ton. Wash. Dep. Fish Wildlife Tech. Rep. H95-08:68 p. Varnavskiy, V. S., and N. V. Varnavskiy. 1985. Assessment of straying between groups of early spawn- ing sockeye salmon, Oncorhynchus nerka. in Nakhichinskoye Lake, Kamchatka. .J. Ichthyol. 25:136-139. Whitman, R. P., T. P. Quinn, and E. L. Brannon. 1982. Influence of suspended volcanic ash on homing be- havior of adult chinook salmon. Trans. Am. Fish. Soc. 111:63-69. 717 Oceanic feeding habits of chinook salmon, Oncorhynchus tshawytscha, off northern California Sharon L. Hunt Timothy J. Mulligan Department of Fisheries Humboldt State University, Areata, California 95521 E-mail address (for T J Mulligan, contact author): t|m2@axe humboldt edu Kenichiro Komori School of Biological Sciences Macquarie University, New South Wales 2109, Australia The chinook salmon, Oncorhynchus tshawytscha, is an important com- mercial and recreational species inhabiting rivers and nearshore coastal waters from San Diego, California, to the Bering Sea and Japan (Miller and Lea, 1972). Many West Coast populations are in a serious decline (Pearcy, 1992). Nehlsen et al. (1991) reported an overall decrease in salmonid num- bers in the coastal waters of the Pacific Northwest and suggested that northern California chinook salmon runs may be at high risk of extinction owing to 1 ) habitat dam- age and mainstream passage prob- lems; 2) overharvesting: and 3) hy- bridization, predation, competition, disease, and poor ocean survival conditions. Surprisingly, little re- search has been done off the north- ern California coast regarding the diet of salmonids during oceanic migrations. Studies on the feeding habits of adult chinook salmon have been conducted from San Francisco, California, to southeastern Alaska. Northern anchovy (EngrauUs mor- (iax), juvenile rockfishes iSebastes spp. ), euphausiids. Pacific herring (Clupea pallasii), osmerids, and crab megalopae (Cancer magister) have been reported as main prey items of chinook salmon ranging from San Francisco to the Washing- ton coast (Heg and Van Hyning, 1951; Merkel, 1957; Petrovich, 1970;Brodeuretal., 1987). Various studies conducted in more northern regions of the Eastern Pacific Ocean have shown Pacific herring and Pacific sandlance (Ammodytes hexapteriis) as dominant food items (Pritchard and Tester, 1944; Reid, 1961; Prakash, 1962). All of the above studies noted seasonal and or annual differences in the dominant prey items. Overall, it appears that northern anchovy and rockfishes are the most important prey items for chinook salmon in southern coastal regions (i.e. San Francisco Bay area) whereas the importance of Pacific herring and Pacific sand- lance increases in more northern regions (Healey, 1991). To ad- equately describe the trophic re- sources utilized by a fish popula- tion, it is necessary to sample at consistent times throughout the year (Bowen, 1996). Furthermore, in an upwelling zone, such as north- ern California, the food habits of many fish species may fluctuate considerably between years owing to environmental variability (Brodeur and Pearcy, 1992). For example, di- ets of pelagic nekton may vary ow- ing to changes in oceanographic con- ditions, such as onshore and along- shore transport, primary productiv- ity, and prey abundances (Brodeur et al., 1987). In our study, we exam- ined the diet of chinook salmon off northern California. The main objec- tives were 1 ) to compare the diet be- tween two consecutive years and 2) to examine seasonal variation in the prey items consumed. Materials and methods Stomach samples were collected from fish caught in coastal waters off Humboldt Bay (40°46'N, 124° 14'W), Trinidad Bay (41°03'N, 124°09'W), and Crescent City (41°46'N, 124°13'W), California (Fig. 1). From May through Sep- tember 1994, 196 stomachs were collected from Chinook salmon from the three areas. During 1995, 112 stomachs were collected from the same ports but only in June and September. Approximately 60*^, 10%, and 30% of the total stomachs collected were taken from Hum- boldt Bay, Trinidad Bay, and Cres- cent City, California, respectively, in both years. Stomach collections were obtained from California De- partment of Fish and Game (CDF &G) port samplers and directly from sportfishermen. Owing to changing and sporadic season opening dates, as well as to varied placement of port samplers and weather conditions, an unbiased random sampling scheme was not possible. All fish were ob- tained fi'om the recreational fishery and were greater than or equal to 22 inches (>56 cm), the CDF&G mini- mum size limit. Total length mea- surements were obtained for 54 fish in 1994 and ranged from 59 to 96 cm (.v=74 cm). Although no fish were measured in 1995, they were simi- lar in size to those collected in 1994. This finding suggests that all fish sampled were three to five years of age (Healey, 1991). Manuscript accepted 23 October 1998. Fish. Bull. 97:717-721 (1999). Fishery Bulletin 97(3), 1999 Chinook salmon stomach contents were removed from the digestive tract, fixed in 109f formaUn, trans- ferred to 40% isopropyl alcohol, and sorted under a dis- secting microscope into major taxonomic groups. When possible, stomach contents were identified to species. The best measure of dietary importance is one where both the number and weight of a food category are recorded (Hyslop, 1980) . Stomach content data were summarized by four methods: 1) numerical, 2) gravimetric, 3) frequency of occurrence, and 4) in- dex of importance. Each food group was enumerated and weighed from each stomach. Wet weight of prey items was measured to the nearest 0.01 g. By using these data, percentage of total number i'/cN), per- centage of total weight (%W), and percentage fre- quency of occurrence C^'AF) were calculated for each food group. "Index of importance" (lOI) was calcu- lated for each food group as follows: lOI. 100 X HI ^, I." HI where ///, 17 7(F + 7(W for food gi'oup a; and the number of different food gi'oups (Hannah, 1980; Grav et al., 1997). Figure 1 Humlxildl Bav, Trinidad liav. and Crescent City. California. Because this index calculation is based on %F as well as V( W, the bias towards heavier, infrequently found prey items is reduced. Results Values for %N, %F, % W, and lOI for prey items en- countered in the stomach analyses are listed in Table 1. All stomachs examined (308) contained food items except one from 1994 and two from 1995. Each prey item was present in stomachs sampled from both 1994 and 1995 except for octopi iOctopus riiheNcens) (2), jacksmelt (Atherinopsis callfornlensis) ( 1 ), cottids (1), pleuronectids (1), Pacific sandlance (108), and rockfishes (45), which were observed only in 1994 samples and isopods (3) which were present only in 1995 samples. Total values for %F did not equal 100% owing to unidentifiable prey items in the diet, espe- cially in 1995 (Table 1). In 1994, the lOI indicated that euphausiids were the predominant food item, accounting for over 27%- of the total. Euphausiids not only ranked highest by %N and %F but also were the leading prey item by % W. In addition, notable lOI values were observed for crab megalopae. Pacific herring, surf smelt (Hypomesiis pretiosiis). Pacific sandlance, northern anchovy, night smelt (Spii-inchus starksi). and squid {Loligo opaleHceus). Infrequently encountered prey items included Pacific saury iCololabis saira), rock- fishes, amphipods, jacksmelt, octopi, shrimp (mysid), juvenile pleuronectids, and juvenile cottids. In 1995, chinook consumed primarily northern anchovy, which represented over 33% of the total lOI but also preyed upon Pacific herring, squid. Pacific saury, surf smelt, night smelt, euphausiids, and crab megalopae. Only rarely were amphipods, isopods, and shrimp found in stomachs. Large interannual variations in lOI can be seen for euphausiids, crab megalopae. Pacific sandlance, northern anchovy, squid, and Pacific saury. Seasonal variation of dominant prey items, for 1994 and 1995, is illustrated in Figure 2, A and B, respectively. In 1994, 88 stomachs were examined from May and June (late spring). lOI values for eu- phausiids (34%), crab megalopae i259r }, and Pacific herring (17% ) dominated all other prey items. Late summer lOI values ( based on 108 stomachs acquired in August and September 1994) indicated Pacific sandlance (22*"^ ), surf smelt (21%), northern anchovy ( 17% ), and euphausiids ( 14% ) to be major prey items. In 1995, 26 stomachs were examined from fish col- lected in June and 86 from September. The lOI val- ues in our study showed that squid (45% ), surf smelt (25%), euphausiids (19%), and Pacific herring ( 11%) are important prey items in late spring whereas NOTE Hunt et al.; Oceanic feeding habits of Oncorhynchus tshawytscha 719 Table 1 Values for percentages by number (%N), frequency (%F) and weight {9c W) for prey items observed in stomachs of chinook salmon. Oncorhynchus tshawytscha, collected in the coastal waters off Hum boldt Bay Tr nidad Bay, and Crescent City, CA, durir g the summers of 1994 and 1995. 1994 1995 %N %F %W lOI %N %F %W lOI Fishes Clupea pallasii 0.1 5.6 17.5 12.2 0.5 5.5 17.8 15.7 Engrail lis mordax 0.1 3.6 8.7 6.5 8.3 14.6 34.6 33.2 Hypumesus prctwsus 0.1 6.7 12.2 10.0 0.4 2.7 9.6 8.3 Spirinchus starksi 0.2 6.7 4.7 6.0 1.3 2.7 7.0 6.5 Cololabis saira 0.0 1.0 4.3 2.8 1.9 4.6 12.7 11.6 Atherinopsis californiensis 0.0 0.5 0.6 0.6 — — — — Ammodytes hexapterus 0.4 8.7 9.8 9.8 — — — — Sebastes spp. (juvenile)^ 0.2 3.6 1.6 2.7 — — — — Cottidae 0.0 0.5 0.0 0.3 — — — — Pleuronectidae 0.0 0.5 0.0 0.3 — — — — Euphausiid Thysanoessa spinifera 71.4 23.1 28.7 27.4 80.8 3.6 5.3 6.0 Crustaceans Cancer magister (megalopa) 27.5 19.5 8.6 14.9 2.7 5.5 0.1 3.8 Mysid 0.0 1.0 0.0 0.6 0.1 0.9 0.0 0.6 Amphipoda Atylus tridens 0.0 2.1 0.0 1.1 2.6 0.9 0.0 0.6 Isopoda Tecticeps convexus 1 ,. , „ . , , - , t combined Synidotea bicuspida J " " 0.2 0.9 0.0 0.6 Cephalopods Loligo opalescens 0.1 5.1 3.2 4.4 1.2 6.4 12.9 13.0 Octopus rubescens 0.0 1.0 0.0 0.6 — — — — Total 100.0 89.2 100.0 100.0 100.0 48.2 100.0 100.0 ' Sebastes entomelas, S. mystinus. S. melanops. S. pinniger iin rank order of abundance! northern anchovy {49'7c). Pacific herring (19^7^), Pa- cific saury ( 17% ), and night smelt ( lOVc ) dominate in late summer. Discussion The feeding habits of northern California chinook salmon varied between years and season. This re- sult is consistent with previous studies. Petrovich (1970) found northern anchovy, euphausiids, Pacific herring, osmerids, and rockfishes to be the most im- portant prey items in coastal waters off Humboldt Bay and Trinidad Bay in 1960-1964. As in our study, crab megalopae were especially abundant in one year In contrast to our results, Petrovich reported rock- fishes to be present in 13% of all stomachs from 1960 to 1964, and in terms of frequency of occurrence they ranked third to northern anchovy and euphausiids. We encountered only 45 juvenile rockfishes during our study, all of which came from seven stomachs in the spring of 1994, representing only 2% of all stom- achs examined. Petrovich found relatively few Pacific sandlance and squid, and no Pacific saury in salmon stomachs that he examined. Our study shows that each of these food items contributes significantly to the diet of chinook salmon off Northern California (Fig. 2). Pa- cific sandlance, although absent in 1995 and only present in small numbers during late spring of 1994, were the major food source (707=22%) in late sum- mer of 1994. Squid were present in late spring and late summer of both years and greatly outranked (707=45% ) all prey items in late spring 1995. Pacific saury were present in stomachs collected during late summer 1994 (707=8%) and late summer 1995 720 Fishery Bulletin 97(3), 1999 ■ Late spring 1994 B Late summer 1994 50 45 40 35 30 25 20 15 10 5 0 B ■ Late spring 1995 BLate summer 1995 L I i li 11 tl -li I 8 -31 a I fr X W 1 2 5 > 13 Figure 2 (A) 1994 late spring and late summer index of importance (/O/i values for predominant prey items consumed by chinook salmon, Oncorhynchus tshawytscha. in the coastal waters off Humboldt Bay, Trinidad Bay, and Crescent City, California. (B) 1995 late spring and late sum- mer index of importance I/O/) values for predominant prey items consumed by chinook salmon, Oncorhynchus tshawytschn. in the coastal waters off Humboldt Bay, Ti-inidad Bay. and Cres- cent City. California. (.I0I=n7c}. The index of importance indicates that interannual and seasonal variation in the feeding habits of chinook salmon does in fact occur, most likely, in response to variation in the relative abun- dance of prey along the coast. Healey (1991) summarized regional trends for coastwide data on chinook feeding habits and recog- nized that the importance of Pacific herring and Pa- cific sandlance increased from south to north whereas the importance of rockfishes and northern anchovy decreased. Our data agree with this summary. The diet of chinook salmon off northern California con- tained characteristic prey items important to chinook found both to the north and south. Although our study NOTE Hunt et al.: Oceanic feeding habits of Oncorhynchus tshawytscha 721 found rockfishes not to be a major prey item of chinook salmon, northern anchovy were in fact com- monly taken. In addition, the importance of Pacific herring and Pacific sandlance was substantial. The occurrence of invertebrate prey items (euphausiids, crab megalopae, and squid) during spring months agrees with studies conducted both to the south (Merkel, 1957) and north (Reid, 1961; Prakash, 1962; Brodeur et al., 1987 ) of the region of our study. Merkel ( 1957 ) commonly found euphausiids and squid, April through June, and crab megalopae (presumed to be Cancer magister). March through May, in stomachs of fish caught off San Francisco. Similarly, Brodeur et al. (1987), in coastal waters off Oregon and Wash- ington, and Prakash (1962), in British Columbian waters, found euphausiids to be a major prey item during the months of May and June. Prakash also observed an abundance of crab megalopae in chinook diet during June. Squid were a dominant prey item one year in Reid's study off southeastern Alaska dur- ing late spring. These results and the great impor- tance of euphausiids, crab megalopae, and squid in our study (Fig. 2) confirm the importance of inverte- brate prey, in some years, to chinook salmon. Acknowledgments Financial support for this research was provided by the U.S. National Marine Fisheries Service. An ear- lier version of this manuscript was critically reviewed by R. D. Brodeur. Literature cited Bowen , S. H. 199G. Quantitative description of the diet, /n B. Murphy and D. Willis (eds.). Fisheries techniques, p. 513-532. Am. Fish. See. Bethesda, MD, 732 p. Brodeur, R. D., H. V. Lorz, and W. G. Pearcy. 1987. Food habits and dietary variability of pelagic nekton off Oregon and Washington. 1979-1984. U.S. Dep. Commer.. NOAA Tech. Rep. NMFS 57, 32 p. Brodeur, R.D., and W.G. Pearcy. 1992. Effects of environmental variability on trophic inter- actions and food web structure in a pelagic upwelling ecosystem. Mar Ecol. Prog. Ser. Vol. 84:101-119. Gray, A. E., T. J. Mulligan, and R. W. Hannah. 1997. Food habits, occurrence, and population structure of the bat ray. MyUobatis californica. in Humboldt Bay. CA. Environ. Biol. Fish. 49:227-238. Hannah, R. 1980. Age, growth and food habits of the small mouth bass, Micropterus dolomieui, in Clair Eagle Reservoir, CA. Unpubl. M.S. thesis. Humboldt State University. Areata, CA, 57 p. Healey, M. C. 1991. Life history of chinook salmon. In C. Groot and L. Margolis, (eds.). Pacific salmon life histories, p. 311- 393. Univ. British Columbia Press. Vancouver, Canada, 564 p. Heg, R., and J. Van Hyning. 1951. Food of chinook and silver salmon taken off the Or- egon Coast. Oregon Fish Commission Research Briefs 3(2):32-40. Hyslop, E. J. 1980. Stomach content analysis-a review of methods and their applications. J. Fish Biol. 17:411-429. Merkel, T. J. 1957. Food habits of king salmon, Oncorhynchus tshawytscha (Walbaum). in the vicinity of San Francisco, California. California Fish Game 43(4):249-270. Miller , D. J., and R. N. Lea. 1972. Guide to coastal marine fishes of California. Calif Dep. Fhsh Game Fish Bull. 157:1-249 Nehlsen, W., J. E. Williams, and J. A. Lichatowich. 1991. Pacific salmon at the crossroads: stocks at risk from California, Oregon, Idaho, and Washington. Fisheries 16(2):4-2L Pearcy, W. G. 1992. Ecology of North Pacific salmonids. Univ. Washing- ton Press. Seattle, WA. 179 p. Petrovich, A., Jr. 1970. Biota of near-shore waters off Humboldt Bay and Trinidad Head. 1960-1964. as shown by diet of Pacific Salmon. Unpubl. M.S. thesis. Humboldt State University, Areata, CA. 69 p. Prakash, A. 1962. Seasonal changes in feeding of coho and chinook (spring) salmon in southern British Columbia waters. J. Fish. Res. Board Can. 19:851-866. Pritchard, A. L., and A. L. Tester. 1944. Food of spring and coho salmon in British Columbia. Bull. Fish. Res. Board Can. 65. 23 p. Reid, G. 1961. Stomach content analysis of troll caught king and coho salmon, southeastern Alaska. 1957-1958. U.S. Fish Wildl. Serv. Spec. Sci. Rep. Fish. 379. 8 p. 722 The development of the digestive tract and eye in larval walleye pollock, Theragra chalcogramma* Steven M. Porter Gail H. Theilacker Alaska Fisheries Science Center National Marine Fisheries Service, NOAA 7600 Sand Point Way NE, Seattle, Washington 981 15 E-mail address (for S M Porter): sportertSafscnoaaaov First feeding ( FF) is a time at which a fish larva must initiate feeding or face starvation that will weaken it and eventually lead to its death. In most oviparous fish with pelagic eggs, organ systems develop that allow larvae to switch from an in- ternal nutrient source (yolk) to an external source (prey) during the time between hatching and FF. Star- vation may be a major cause of the high mortalities that occur during the larval period and thus may af- fect recnaitment ( Hjort, 1914; O'Con- nell, 1980; Theilacker, 1986). Indeed, some studies suggest that reci-uit- ment is determined in this period (Houde, 1987;Freebergetal., 1990). Laboratory experiments show that at FF, the gi'owth rate of walleye pol- lock, Theragra chalcogramma, lar- vae decreases (Yamashita and Bailey, 1989), or even ceases for a period (Theilacker and Shen, 1993). Fur- thermore, delaying the introduc- tion of food at FF causes a reduc- tion of the gi'owth rate and size-at- age (Theilacker and Shen, 19931. These studies indicate that FF walleye pollock lack energy re- serves, and thus the availability of nutritious prey at this time is criti- cal. Theilacker and Porter (1995) found that the largest proportion of starving walleye pollock larvae in Shelikof Strait, Gulf of Alaska, are in the size class that includes FF larvae. Thus, the FF period appears to be the time when walleye pollock larvae are most vulnerable to star- vation. Studies involving larvae of other marine fish have found simi- lar results (Atlantic mackerel. Scomber ,sco;?;6/7/.s. Ware and Lam- bert, 1985; jack mackerel, Tixichurus sytiimetricus, Theilacker, 1986). Larval walleye pollock survival is dependent upon timely develop- ment of organs required for feed- ing. Of the many developmental changes occurring during the early larval period, two of the most im- portant for walleye pollock may be vision and the digestive tract. Eye- sight is probably the most impor- tant sense walleye pollock larvae use in finding prey because they are visual predators (Paul, 1983) and do not use chemosensory cues (Davis and 011a, 1995). Digestive tract de- velopment is important for efficient assimilation of nutrients needed for growth. In this study, histological sections were used to describe the development of the mouth, diges- tive tract, and eyes in laboratory- reared walleye pollock, from hatch- ing to 31 days after hatching (DAH), to examine how organs nec- essary for feeding develop. This is essential to an understanding of why walleye pollock larvae are most vulnerable to starvation dur- ing the first week of feeding. Materials and methods In 1991 and 1992, adult walleye pollock were collected from Shelikof Strait, Gulf of Alaska, and spawned aboard ship. Fertilized eggs were maintained aboard ship at 3°C for a few days, then transported to the laboratory. In 1991, walleye pollock larvae were reared at Friday Har- bor Laboratories, University of Washington, San Juan Island, Washington, and in 1992, at the Alaska Fisheries Science Center, Seattle, Washington. Between 500 and 1000 larvae were reared in a black, circular, 120-L fiberglass tank (62 cm diameter, 43 cm deep) filled with 90 L filtered seawater (salinity=28.0 ±0.5 ppt) and main- tained at 6° ±0.5"C, the typical sea- water temperature in Shelikof Strait when walleye pollock larvae initiate feeding (Kendall et al., 1987). Larval rearing procedures followed Porter and Theilacker 1996.1 Fluorescent fixtures were used with a 16-hour light cycle; the amount of light at the surface of the water in the rearing tank was 17 |i mol photon/m-/s. Pi-ey consisted of the rotifer Brachionus plicatilis, raised on an algal diet oflsochrysis galbana and Pavlova lutheri high in unsaturated fatty acids (Nichols et al., 1989), as well as of Acartia sp. copepod nauplii and copepodites collected from a local lagoon. At 3 DAH, four to five days before FF, rotifers were added at 10/mL and Acartia were added at 3/mL to the rearing tank and maintained at this level throughout the rearing ■' Contribution 0285-RAO-O to Fisheries Oceanogi'aphy Coordinated Investigations (FOCI I, NOAA, 7600 Sand Point Way NE, Seattle, WA 98115. ' Porter, S. M., and G. H. Theilacker 1996. Larval walleye pollock, Theragra chalco- gramma. rearing techniques used at the Alaska Fisheries Science Center. Seattle Washington. AFSC processed report 96- 06, 26 p. U.S. Dep, Commerce, NOAA, NMFS, Alaska Fisheries Science Center, 7600 Sand Point Way NE, Seattle, WA., 98115. Manuscript accepted 17 July 1998. Fi.sh. Bull. 97:722-729 (1999). NOTE Porter and Theilacker: Development of digestive tract and eye in larval Theragra chalcogramma 723 period. Eight to 10 larvae were sampled every day or every other day after hatch- ing to 23 DAH in 1991 and to 31 DAH in 1992. Larvae were placed into either Bouin's fixative which was replaced with VO'/r ethanol 24 to 48 h later, or Z-fix^' (a solution of 10% formalin with zinc and buffers added). Larvae were dehydrated in a graded series of butyl alcohols, em- bedded in paraffin wax, sectioned sagit- tally into 6-|im serial sections, and stained with hematoxylin and eosin (H and E). Standard lengths (SL, tip of upper jaw to end of notocord) of preserved larvae were measured to the nearest 0.08 mm and ad- justed to live SL (Theilacker and Porter, 1995). To describe development of the mouth, digestive tract, and eyes, histologi- cal sections of 5-10 larvae were examined from each sampling day. Results Larval growth In 1991, the mean SL at hatching was 4.27 ±0.07mm, whereas in 1992 hatching size averaged 4.66 ±0.15mm (Fig. 1). The de- crease in average size of larvae sampled on day 23 in 1991 (Fig. 1) was probably due to the sampling of slow growing lar- vae in poor condition. A bowl was used to sample larvae from the tank, and we sus- pect that for this sample only slow grow- ing larvae were captured. The overall mean growth rate from hatching to 20 DAH was 0.12 mm/d in both 1991 and 1992. In 1992, experiments continued un- til 31 DAH and average growth from hatching was 0.10 mm/d. The onset of FF occurred over a period of three days; day of first feeding was defined as the time when 50% of the larvae began feeding. In 1991, day of first feeding occuired at 8 DAH, and the mean SL was 5.82 ±0.12 mm. In 1992, larvae began feeding at 7 DAH, and the mean SL was 5.73 ±0.25 mm (Fig. 1). Mouth and digestive tract development At hatching a membrane covered the mouth, and jaw cartilage had not yet formed. At 2 DAH the mem- Days after hatching Figure 1 Growth of walleye pollock, Theragra chalcogramma. reared at 6'C in the laboratory from hatching to 2.3 days after hatching in 1991, and to 31 days after hatching in 1992. Mean standard length (preserved length adjusted to live length, see text) and standard deviation are shown; 5 to 10 larvae were measured at each age. The small size for 1991 at 23 days after hatching may be due to sampling slow growing larvae in poor condition (see text). Arrow indicates day of first feeding. 2 Anatech, Ltd., Battle Creek, MI. brane began to degenerate leaving an orifice that was 22 to 55 |jni wide. Mouth and jaw cartilage also be- gan forming at this time, and the cartilage in the roof of the mouth, the trabeculum cranii, became most distinct. All jaw cartilage elements were present and almost completely formed at 5 DAH (approxi- mately 2 days before FF); the mouth was considered functional 5 to 6 DAH (Table 1). At hatching the gut was a straight tube with a narrow lumen. The foregut could be distinguished 724 Fishery Bulletin 97(3), 1999 Table 1 Mean standard length (tip of upperjaw to end of notoeord) and age at the development of various elements of the mouth, digestive tract, and eye for walleye pollock, Theragra chalcogramma. reared at 6'C in the laboratory in 1991 and 1992. '+', vacuoles were not observed at 23 DAH in 1991, the last sampling day. Preserved standard length adjusted to live standard length (Theilacker and Porter, 1995). SD = standard deviation, n = number of larvae measured. Age = days after hatching. Mean standard length mm (SD) 1991 1992 Age (year) Mouth and digestive tract Foregut. midgut, and hindgut separated by valves Jaw developed and mouth functional Folding of midgut epithelium Midgut coiling, eosinic vesicles in the hindgut Lipid vacuoles in midgut Eye Ocular motor muscles apparent Eye fully pigmented Lens retractor muscle developed 5.15(0.15) 5 5.18(0.30) 5 3(1991,92) 5.56(0.28) 5 5.65(0.08) 5 5(1992) 6(1991) 6.19(0.25) 5 6.14(0.22) 10 11(1992) 12(1991) 6.14(0.28) 5 6.64(0.15) 8 13(1992) 14(1991) + 7.49(0.54) 10 23(1992) 5.15(0.15) 5 5.18(0.30) 5 3(1991,92) 5.38(0.13) 5 5.48(0.21) 5 4(1991,92) 6.22(0.331 5 no samples taken 15(1991) by its cuboidal epithelium, but no distinction could be made between the midgut and hindgut which were lined with similar columnar epithelium. At 1 DAH, the gut lumen widened and began to separate into a long midgut and short hindgut that were demarcated by a constriction at the future site of the ileocaecal valve. At 3 DAH, the pyloric valve separated the foregut and midgut, and the ileocaecal valve separated the midgut and hindgut (Table 1; Fig. 2). At 11 to 12 DAH (4 days after FF) the gut began to develop large folds, and be- gan to coil 13 to 14 DAH (6 days after FFt (Table 1). Also, at 13 to 14 DAH, eosinic vesicles (also referred to as eosinophilic granules or inclusions), ranging in size from 1 to 3 |.im, began appearing in the apical cyto- plasm of the hindgut epithelial cells (Table 1). Vacu- oles in midgut epithelial cells were first observed at 23 DAH in 1992 larvae (Table 1). The vacuoles were not distinct and observed only in one larva out often. Wall- eye pollock larvae showed no other changes in gut sti-uc- ture (i.e. no stomach, or pyloric caeca) up to 31 DAH, Eye development At hatching, a lens was present, and small scattered patches of pigment were located throughout the retina ( Fig. 3 ). The pigment patches increased in size and joined to form the pigment layer; by 4 DAH the eye was fully pigmented (Table 1). Ocular motor muscles were not developed at hatching, but at 3 DAH these muscles became apparent (Table 1). At 9 DAH, the lens retractor muscle was identified as a thin stioicture that connected the lens to the retina. From 9 to 15 DAH the muscle increased in width as it grew. The size and shape of this muscle changed very little beyond 15 DAH; therefore it was considered func- tional beginning at this time (Fig. 4, Table 1). Discussion Larval growth Growth of walleye pollock larvae in this study was similar between years (0.12 mm/d from hatching to 20 DAH for both 1991 and 1992) and to rates found by others for walleye pollock larvae reared at 6°C in the laboratory; 0.14 mm/d from hatching to 19 DAH (Theilacker and Shen, 1993), 0.11 mm/d from hatch- ing to 21 DAH (Yamashita and Bailey, 1989) and 0.065 mm/d from hatching to 15 DAH (Nishimura and Yamada, 1984). Growth was slightly lower than the 0.14 to 0.23 mm/d calculated for field-collected walleye pollock larvae (Bailey et al., 1996). Mouth and digestive tract From observations on live larvae, Bailey and Stehr ( 1986) found that walleye pollock have no mouth at hatching; the mouth begins to develop 2 days later NOTE Porter and Theilacker: Development of digestive tract and eye in larval Theragra chakogramma 725 Figure 2 Walleye pollock. Theragra chalcogramma. larva 3 days after hatching. The foregut (FG) and midgut (MG), and midgut and hindgut (HGl are separated by the pyloric and ileocaecal valves respectively (arrowheads I. (Bouin's fixative, H and E; standard length=5.24 mm: scale bar=100 |ami. Y = yolk. Figure 3 Walleye pollock, Theragra chalcogramma. eye at hatching. The eye is not fully pig- mented. Arrowheads identify patches of pigment. (Bouin's fixative, H and E; standard length=4.34 mm; scale bar=50 |am). L = lens. 726 Fishery Bulletin 97(3), 1999 Figure 4 The lens retractor muscle (LR) in the eye oi' a 15-day posthatching walleye pollock, Theragra chalcogramma. larva. (Bouin's fixative, H and E; standard length=6.65 mm; scale bar=20 uml. L = lens. and is functional by 4 to 5 DAH. Histological analy- sis in this study showed that mouth and jaw devel- opment proceeded at a similar rate; the jaw was fully formed and the mouth was functional 5 to 6 DAH. As in Atlantic cod, Gadus morhua, larvae (Kjorsvik et al., 1991), the first feature of the walleye pollock mouth to form was the trabeculum cranii ("roof of the mouth"). At hatching, walleye pollock larvae had a straight- tube gut. This simple gut arrangement has been noted for larvae of other species offish as well (north- ern anchovy, Engraulis mordax, O'Connell, 1981; Atlantic cod, G. morhua, Kj0rsvik et al., 1991; tur- bot, Scophthalinus maximus, Segner et al., 1994). At 3 DAH, three distinct portions of the gut could be identified: the foregut, midgut, and hindgut. This gut arrangement is typical of larval fish (Govoni et al., 1986) and walleye pollock larvae showed no major changes in gut structure (i.e. no stomach, or pyloric caeca) up to 31 DAH. For larval fish, major changes in gut structure happen rapidly at metamorphosis rather than gradually during the larval period (Govoni et al., 1986). However, at 11 to 12 DAH (4 days after FF) the larval walleye pollock midgut epi- thelium began to develop large folds, and the gut began to coil 13 to 14 DAH (6 days after FF). For walleye pollock larvae reared at 6''C, Oozeki and Bailey ( 1995 ) noted that gut coiling began on 16 DAH and was complete by day 23. Midgut coiling increases the length of the gut, and the residence time of prey in it ( Yamashita and Bailey, 1989). At one week after FF, both epithelial folds and gut coiling enable wall- eye pollock larvae to assimilate nutrients more effi- ciently through increased absorptive surface area and longer residence times. Histological evidence suggests the larval walleye pollock gut functions in the same manner as in other fish larvae. The midgut of fish larvae digests and absorbs lipids (Govoni et al., 1986) and this appears to be the function of the larval walleye pollock mid- gut as well. For larval whitefish, Coregonus fera , lipid vacuoles appear as circular "voids" in the apical por- tion of the midgut epithelial cells one day after feed- ing (Loewe and Eckmann, 1988). In other species of fish, similar vacuoles are also reported to contain lipid (goldfish, Carassius aiiratus, Iwai, 1968; tur- bot, S. t7w.xlmiis. Segner et al., 1994). In our study, vacuoles (circular voids) were observed in midgut epithelial cells of one walleye pollock larva at 23 DAH. This is later in development than they have been observed in other fish larvae but lipase is present in walleye pollock larvae at hatching; there- fore lipid digestion could occur at FF. Lipase activity increases with age (Oozeki and Bailey, 1995) so that as larvae grow, more lipid can be digested and this could produce larger, more visible (by H and E stain- NOTE Porter and Theilacker: Development of digestive tract and eye in larval Theragra chalcogramma 727 ing) vacuoles with age. There was probably lipid in the midgut cells earlier than 23 DAH but in such small amounts so that it could not be identified with- out special staining. The larval walleye pollock hind- gut appears to function in protein digestion like the hindgut of other species offish larvae (Govoni et al., 1986). Watanabe (1984) demonstrated that protein is pinocytotically moved into hindgut epithelial cells for intracellular digestion. Iwai and Tanaka (1968) stated that the intracellular protein appears as eosi- nophilic granules (also referred to as eosinophilic vesicles or inclusions) within the apical portion of the hindgut cells. At 13 to 14 DAH (5 to 7 days after FF), eosinic vesicles appear in the apical cytoplasm of the hindgut epithelial cells of walleye pollock lar- vae. For Atlantic cod, G. morhiia, larvae (Kjorsvik et al., 1991), a close relative of walleye pollock, inclu- sions were not observed until 2 to 5 days after FF. In other studies eosinophilic inclusions have been ob- served soon after FF (northern anchovy, E. mordax, O'Connell, 1981; whitefish, C. fera, Loewe and Eckmann, 1988; turbot, S. maximus. Segner et al., 1994). Trypsin is present in walleye pollock larvae at hatching, its activity increases with age, and it is not supplemented by prey in the gut (Oozeki and Bailey, 1995). Eosinophilic vesicles appeared at the time the gut began to coil, and both coiling and incre- ased trypsin would allow increased digestion of prey making more protein available to the hindgut, possi- bly explaining why vesicles were not apparent until some time after FF. The appearance of lipid vacuoles and eosinophilic vesicles after the first week of feeding provides evidence for improved digestive capability. Eye Sight is probably the most important sense walleye pollock larvae use for feeding. Walleye pollock rely on vision to search for prey (Paul, 1983) and do not respond to chemosensory cues (Davis and 011a, 1995). Laboratory experiments show that a group of wall- eye pollock larvae, age 21 DAH, remains aggregated when rotifers are introduced into the group, but when only the scent of rotifers is introduced, the group dis- perses (Davis and 011a, 1995). Because they are vi- sual predators and because light below 0.006 )i mol photon/m^/s limits their ability to capture prey (Paul, 1983), walleye pollock larvae probably have a pure- cone retina like many other species of fish larvae. Blaxter and Staines (1970) examined the retina of 12 species offish larvae including haddock, Melano- gramnuis aeglefinus (which belongs to the same fam- ily as walleye pollock, Gadidae), they showed that had- dock and seven other species have a pure cone retina. 011a and Davis ( 1990) stated that the retina of walleye pollock larvae most likely does not contain rods. It is unknown when rods begin to appear in walleye pol- lock, but for herring, Clupea harengus (Blaxter and Jones, 1967), plaice, Pleuronectes platessa (Blaxter, 1968), and sole, Solea solea (Sandy and Blaxter, 1980), rods begin to appear at metamorphosis. The lack of eye pigmentation at hatching has been found for many teleosts (Blaxter, 1986), including walleye pollock (Bailey and Stehr, 1986; our study), and eyes are probably nonfunctional at this time (Blaxter, 1986). Ocular motor muscles developed just as walleye pollock eyes became fully pigmented (3 DAH). The development of eye pigmentation and ocular motor muscles should provide a FF larva with sight and the ability to move its eyes. The lens re- tractor muscle did not develop until after FF; it was considered functional beginning at 15 DAH. The lens retractor muscle develops after FF in other species of fish as well (northern anchovy, E. moi'dax, OConnell, 1981; white seabass, A^rac^osczo/! nobilis, Margulies, 1989). This muscle allows a larva to fo- cus on objects at different distances, thereby incre- asing the field of vison (Munz, 1971). Because this muscle was not functional until 15 DAH, for about a week after FF a larval walleye pollock's field of view is restricted because it is unable to change the focus of its eyes. Thereafter, visual acuity improves, allow- ing walleye pollock larvae to detect both prey and predators more easily. In Shelikof Strait, Gulf of Alaska, the proportion of starving walleye pollock larvae decreases dramati- cally after the first week of feeding (Theilacker and Porter, 1995; Theilacker et al., 1996) and the physi- ological condition of walleye pollock larvae improves as they grow; that is, fewer lai-vae are found in poor condition (Theilacker et al., 1996). Improvements to vision (the lens retractor muscle) and a combination of morphological changes to the gut (folding and coil- ing) as well as an increase in digestive enzymes (Oozeki and Bailey, 1995) contribute to walleye pol- lock larvae becoming less vulnerable to starvation after the first week of feeding. Additional contribut- ing factors may include developmental changes oc- curring to other organ systems (e.g. development of trunk musculature and lateral line system as has been shown for other species offish larvae; Blaxter, 1986; O'Connell, 1981), and larvae becoming better predators as they grow. For northern anchovy, E. mor-dax. larvae, feeding success rapidly improves during the first week of feeding (Hunter, 1972). At hatching walleye pollock larvae lack functional eyes and mouth, and have a straight-tube gut; the devel- opment of these between hatching and FF allows lar- vae to begin feeding, and their continued development after FF improves the larvae's chance of survival. 728 Fishery Bulletin 97(3), 1999 Acknowledgments We would like to thank the following people for their help: Debbie Blood, Ric Brodeur, Jay Clark, Nazila Merati, and Matt Wilson for spawning walleye pol- lock and bringing the eggs back to Seattle; Annette Brown and Stella Spring for assistance rearing the larvae; Frank Morado and Linda Cherepow for pro- viding assistance during the histological phase of this sttidy. Kevin Bailey, Ail Kendall, Mike Canino, Frank Morado, and anonymous reviewers provided helpful comments on drafts of this manuscript. The Univer- sity of Washington, Friday Harbor Laboratory, pro- vided aquarium space in 1991. Literature cited Bailey, K. M., and C. L. Stehr. 1986. Laboratory studies on the early life history of the walleye pollock, Theragra chalcogranima ( Pallas ). J. Exp. Mar. Biol. Ecol. 99:233-246. Bailey, K. M., A. L. Brown, M. M. Yoklavich, and K. L. Meir. 1996. Interannual variability in growth of larval and juve- nile walleye pollock Theragra chalcogramma in the western Gulf of Alaska, 1983-91. Fish. Ocean. .5 (suppl. 11:137-147. Blaxter, J. H. S. 1968. Light intensity, vision, and feeding in young plaice. J. Exp. Mar. Biol. Ecol. 2:293-307. 1986. Development of sense organs and behavior of teleost larvae with special reference to feeding and predator avoidance. Ti-ans. Am. Fish. Soc. 115:98-114. Blaxter, J. H. S., and M. P. Jones. 1967. The development of the retina and retinomotor re- sponses in the herring. J. Mar Biol. Assoc. U. K. 47:677-697. Blaxter, J. H. S., and M. Staines. 1970. Pure-cone retinae and retinomotor responses in lar- val teleosts. J. Mar Biol. Assoc. U. K. 50:449-460. Davis, M. W., and B. L. OUa. 1995. Formation and maintenance of aggregations in wall- eye pollock, Theragra chalcogramma, larvae under labo- ratory conditions: role of visual and chemical stimuli. Environ. Biol. Fish. 44:385-392. Freeberg, M. H., W. W. Taylor, and R. W. Brown. 1990. Effect of egg and larval survival on year-class strength of lake white fish in Grand Traverse Bay, Lake Michigan. Trans. Am. Fish. Soc. 119:92-100. Govoni, J. J., G. W. Boehlert, and Y. Watanabe. 1986. The physiology of digestion in fish larvae. Environ. Biol. Fishes 16:59-77. Hjort, J. 1914. Fluctuations in the great fisheries of northern Eu- rope viewed in the light of biological research. Rapp, P- V. Reun. Cons. Int. Explor. Mer 20:1-228. Houde, E. D. 1987. Fish early life dynamics and recruitment variability. Am. Fish. Soc. Symp. 2:17-29. , Hunter, J. R. 1972. Swimming and feeding behavior of larval anchovy, Engraulis Mordax. Fish. Bull. 70:821-838. Iwai, T. 1968. The comparative study of the digestive tract of te- leost larvae — V. Fat absorption in the gut epithelium of goldfish larvae. Bull. Jpn. Soc. Sci. Fish. 34:973-978. Iwai, T., and M. Tanaka. 1968. The comparative study of the digestive tract of te- leost larvae — III. Epithelial cells in the posterior gut of halfbeak larvae. Bull. Jpn. Soc. Sci. Fish. 34:44-48. Kendall, A. W., Jr., M. E. Clarke, M. M. Yoklavich, and G. W. Boehlert. 1987. Distribution, feeding, and growth of larval walleye pollock, Theragra chalcogramma. from Shelikof Strait, Gulf of Alaska. Fish. Bull. 85:499-521. Kjorsvik, E., T. Van Der Meeren, H. Kryvi, J. Arnfinnson, and P.G. Kvenseth. 1991. Early development of the digestive tract of cod lar- vae, Gadiia morhiia, during start- feeding and starvation. .J. Fish Biol. 38:1-15. Loewe, H,, and R. Eckmann. 1988. The ontogeny of the alimentary tract of coregonid larvae: normal development. J. Fish Biol. 33:841-850. Margulies, D. 1989. Size-specific vulnerability to predation and sensory system development of white seabass, A^ra/o.sc/o;; nobilis, larvae. Fish. Bull. 87:.537-552. Munz, F.W, 1971. Vision: visual pigments. In W. S. Hoar, and D. J. Randall (eds.l. Fish physiology, vol. 5, Sensory systems and electric organs, p. 1-8. Academic Press, New York, NY. Nichols, P, D„ D. G. Holdsworth, J. K. Volkman, M. Daintith, and S. Allanson. 1989. High incorporation of essential fatty acids by the ro- tifer Brachionut; plicatilis fed on the prymnesiophyte Pavlova luthert. J. Mar Freshwater Res. 40:645-655. Nishimura, A., and J. Yamada. 1984. Age and growth of larval and juvenile walleye pol- lock, Theragra chalcogramma (Pallas), as determined by otolith daily growth increments. J. Exp. Mar. Biol. Ecol. 82:191-205. O'Connell, C. P. 1980. Percentage of starving northern anchovy, Engi-aulis mordax. larvae in the sea as estimated by histological methods. P'ish. Bull. 78:475-489. 1981. Development of organ systems in the northern an- chovy, Engraulis niordax, and other teleosts. Am. Zool. 21:429-446. Olla, B. L., and M. W. Davis, 1990. Effects of physical factors on the vertical distribu- tion of larval walleye pollock Theragra chalcograiiima un- der controlled laboratory conditions. Mar. Ecol. Prog. Ser. 63:105-112. Oozeki, Y, and K. M. Bailey. 1995, Ontogenetic development of digestive enzyme activi- ties in larval walleye pollock, T/ieragra clialco.i^i-amma. Mar Biol. 122:177-186. Paul, A. J. 1983. Light, temperature, nauplii concentrations, and prey capture by first feeding pollock larvae Theragra chalco- gramma. Mar Ecol. Prog. .Ser 13:17,5-179. Sandy, J. M„ and J, H, S. Blaxter. 1980. A study of retinal development in larval herring and .sole. J. Mar Biol. Assoc. U.K. 60:59-71. Segner, H., V. Storch, M. Reinecke, W. Kloas, and W. Hanke. 1994. The development of functional digestive and meta- bolic organs in turbot, Scoplttlialmii.'' inaximi/s. Mar. Biol. 119:471-486. Theilacker, G. H. 1986. Starvation-induced mortality of young sea-caught jack NOTE Porter and Theilacker: Development of digestive tract and eye in larval Theragra chalcogramma 729 mackeral, 7>-a(7i(;rus svmmtVn'cus. determined with histologi- cal and morphological methods. Fish. Bull. 84:1-17. Theilacker, G. H., K. M. Bailey, M. F. Canino, and S. M. Porter. 1996. Variations in larval walleye pollock feeding and con- dition: a synthesis. Fish. Ocean. 5 isuppl. 11:112-123. Theilacker, G. H., and S. M. Porter. 1995. Condition of larval walleye pollock, Theragra chalco- gramma, in the western Gulf of Alaska assessed with histo- logical and shrinkage indices. Fish. Bull. 93:333-344. Theilacker, G. H., and W. Shen. 1993. Calibrating starvation-induced stress in larval fish using flow cytometry. Am. Fish. See. Symp. 14:85-94. Ware, D. M., and, T. C. Lambert. 1985. Early life history of Atlantic mackerel ^Scomber scombrus) in the southern Gulf of St. Lawrence. Can. J. Fish. Aquat. Sci. 42:577-592. Watanabe, Y. 1984. An ultrastructural study of intracellular digestion of horseradish peroxidase by the rectal epithelial cells in lar- vae of a freshwater cottid fish Cottus noawae. Bull. Jpn. Soc. Sci. Fish. 50:409-416. Yamashita, Y., and K. M. Bailey. 1989. A laboratory study of the bioenergetics of larval walleye pollock, Theragra chalcogramma. Fish. Bull. 87:525-536. 730 Evaluation of small T-anchor and dart tags for use in marking hatchery-reared juvenile red drum, Sciaenops ocellatus Brent L. Winner Robert H. McMichael Jr. Laurel L. Brant Florida Marine Research Institute Florida Department of Environmental Protection 100 Eighth Avenue SB, St, Petersburg, Florida 33701-5095 Email address (For B L Winner) W/innerBgepic7dep State fl us Red drum, Sciaenops ocellatus, is an estuary-dependent marine spe- cies found in coastal and nearshore waters in the western Atlantic Ocean from Maine to Florida and in the Gulf of Mexico from Florida to Vera Cruz, Mexico (Yokel, 1966, 1980). Red drum are highly sought after as food and gameflsh. Annual commercial landings in Florida av- eraged nearly 1.0 million pounds from the early 1960s until 1986, when the sale of red drum was pro- hibited. Recreational fishing effort directed toward red drum in Florida has more than doubled since 1989 on both the Atlantic and Gulf coasts (Murphy'). Increased fishing pres- sure on red drum stocks has led to the implementation of various man- agement strategies, such as closing the commercial fishery and reducing the recreational catch by means of bag limits, size limits, and closed sea- sons. In the 1970s, methods were developed for culturing red drum in controlled hatchery environments, providing a pathway for the use of stock enhancement for managing the fishery (Arnold et al., 1977; Roberts et al., 1978a, 1978b). Red drum stock enhancement has been conducted ex- tensively in Texas and, to a lesser de- gree, in Florida and South Carolina (Matlock et al., 1986; Willis et al., 1995; Smith etal.^). Monitoring stocked fish popula- tions in the wild is critical in deter- mining the impact of stocked fish on the fishery and the natural population. Red drum are stocked at various sizes: (phase-1 (25-50 mm SL); phase-2 (50-100 mm SL); and phase-3 ( 100-200 mm)). Stock- ing larger size red drum (phase-2 or -3), facilitates the use of tagging as a method of tracking these fish in the wild. Tagged fish can provide valuable data on stock identity, fishing pressure, movements, abun- dance, age and growth, mortality, and stocking-program success (Ricker, 1956; Hilborn et al., 1990; McFarlane et al., 1990). However, the usefulness of these data de- pends on knowledge of tag-reten- tion rates and of the effects of tag- ging on fish growth and survival. A variety of tag types have been tested on red drum of various sizes. Several studies have evaluated the use of coded wire tags in juvenile red drum (phase-2, 50-100 mm SL, Bumguardner et al., 1990, 1992; Szedlmayer and Howe, 1995). Be- cause coded wire tags are not ex- ternally visible, recreational and commercial fishermen are not likely to see them and therefore are not likely to submit capture data to authorities; externally visible tags are better suited for obtaining capture information from recre- ational and commercial fishermen. Externally visible tags are typi- cally restricted to use in red drum larger than 100 mm (phase-3 fish). Retention of externally visible tags in phase-3 red drum ( 100-200 mm SL) has not been rigorously tested, and little is known of the effects of these tags on survival or growth in this size of red drum. Some studies have tested the effects of various external tags on large red drum (>300mmSL;Elam, 1971; Weaver, 1976; Hein and Shepard, 1980a; Gutherz et al., 1990 ). Matlock et al. (1984) reported using Monel jaw tags and Willis et al. (1995) re- ported using internal-anchor tags in fmgerling red drum (100-200 mm SL), but neither report ad- dressed tag-retention rates or ef- fects of tags on growth or survival. The Florida Department of En- vironmental Protection's Stock En- hancement and Research Facility (SERF) has the capability of rear- ing phase-1, -2, and -3 red drum for release. Data from this study were used to evaluate the efficacy of two types of external tags in hatchery- reared phase-3 juvenile red drum, and to provide tag-retention and survival estimates, which are nec- essary for correctly interpreting capture information. Materials and methods Retention of T-anchor and dart tags in hatchery-reared red drum ( 102- 173 mm SL) was evaluated over a 423-day period. Mortality associ- ' Murphy, M. D. 1994. A stock assess- ment for red drum, Sciaenops ocellatus, in Florida. Florida Marine Research Institute. Report to the Fla. Mar. Fish. Comm., Tallahassee, FL, 28 p. •^ Smith T. I. J., M. R. Denson, D. B. White, andW. E.Jenkins. 1993. Evaluation of a preliminary red drum stock enhance- ment program in South Carolina. SC Wildl. Mar Res. Dep., Mar Resour Res. Inst. Annu. Performance Rep. Charleston. SC, 21 p. Manuscript accepted 16 September 1998. Fish. Bull. 97:730-73.5 (1999). NOTE Winner et al : Evaluation of T-anchor and dart tags for use in marking Saaenops ocellatus 731 ated with tagging was monitored for the first 111 days of the experiment. T-anchor tags (lEX tags"^, Fig. 1) had an 18-mm "T" with molded polyethylene on both sides for sup- port. Attached to the "T" portion of the tag was a 42- mm streamer, the distal two-thirds of which was en- cased in polyethylene. T-anchors were inserted through a 1-mm-diameter hole made with a pointed, stainless-steel rod. Tags were inserted on the left ventral side of the fish near the distal end of the pel- vic fin. Once inserted, the tag streamer was pulled so that the T-anchor rested along the body cavity wall, with the streamer protruding. Dart tags (PDX tags-, Fig. 1) had a 9-mm, semi- rigid plastic barb attached to a 45-mm plastic streamer that was encased in polyethylene. Dart tags were inserted into the pterj'giophores of the spinous dorsal fin by using a stainless-steel canula. We dipped all tags and applicators in Betadyne before tagging each fish to minimize the possibility of infection. The streamer of each tag was imprinted with a unique tag number so that we could identify individual specimens. All red drum used in this experiment were spawned and maintained in ponds at the Florida Department of Environmental Protection's Stock Enhancement Research Facihty (SERF) in Port Manatee, Florida. The holding ponds were drained at 290 days after spawning and fish were dip-netted to a small hold- ing tank, where they were anesthetized (Tricaine menthanesulphonate, MS222, 100 ppm). Fish were measured (SL mm) and randomly assigned to one of three treatment groups: dart tagged, T-anchor tagged, and controls (handled but not tagged). Each treatment group contained 100 fish (mean SL=140 mm, mean weight=40 g), divided equally between two 1.5-m-diameter net pens with a volume of 2.12 m'^ Net pens were constructed of 6.25-mm knotless ny- lon mesh and were submerged within a closed aquarium system (>79,000 liters) with supporting aeration, bio-filter, and rapid sand filter. Salinity ( '?<:), dissolved oxygen (mg/L), and temperature (°C) were monitored daily. Ammonia concentration (NH.^^, mg/L) was monitored periodically. Fish were fed a combination of cut squid and commercially prepared pelletized fish food at a rate of 2-5'7f of body weight per day. Net pens were checked daily for fish mortalities and shed tags. Standard length (SL mm), date, tag type, tag number, and pen number were recorded for all fish mortalities. At 66, 111, and 423 days after \ B Figure 1 External tags used to mark hatchery-reared red drum; I A) T-anchor (lEX tags) and (B) dart (PDX tags). Both tags are manufactured by Hallprint Ltd., Holden Hill, Australia. Shown at 75'7< of actual size. ' Hallprint Ltd., 27 Jacobsen Crescent, Holden Hill, South Aus- tralia 5088, Australia. tagging (DAT), fish from all net pens were measured (SL mm), tag numbers checked, and tag wounds ob- served for degree of healing (wound open or closed). Control fish were removed from the study at 1 1 1 DAT. We continued to monitor tagged fish from 111 to 423 DAT so that we could evaluate long-term tag reten- tion. All fish were treated ior Amyloodinium sp. and parasitic copepods at 21 DAT. Standard length measurements of red drum were compared between net pens at 0, 66, and 111 DAT by using a nonparametric Kruskal-Wallis one-way ANOVA on ranks test (a=0.05). Instantaneous sur- vival rates of red drum were compared between treat- ments by using a two-way (% offish alive) repeated measures ANOVA (a=0.05) on arcsine-transformed data. Net pen was used as the subject, with tag type as a group factor and days as a level factor (Fox et al., 1995). Results and discussion Tag-retention rates in red drum in this study were similar for both T-anchor and dart-tagged fish (Table 1 ). Average retention rates for T-anchor and dart tags were, respectively, 98^7^ and 99'A at 66 DAT; 97% and 95% at 111 DAT; 91% and 89% at 423 DAT. No tags were shed until 41 DAT, and nearly all shedding stopped by 150 DAT (Fig. 2). Both tag types were retained in some red drum as large as 485 mm SL. Wallin et al. ( 1997) also reported high short-term (30 days) retention rates (100%) of T-anchor tags in ju- 732 Fishery Bulletin 97(3), 1999 Cumulative number of stied tags • — • T-anchor 1 O T-anchor 2 T-T Dart 1 V— V Dart 2 -y / T ^'^ J7 1 0 50 100 150 200 250 300 350 400 450 Days Figure 2 Cumulative number of tags shed by hatchery-reared red drum over a 42.3-day period. Each tag type was tested in two separate mesh-net pens, each holding 50 fish at start of test. venile Centropomus undecimalis. Collins et al. ( 1994) found that T-anchor tags had higher long-term re- tention rates than did dart tags in juvenile Ac/penser brevirostrum (T-anchor: 92% at 306 DAT; Dart: 509^ at 154 DAT). The low retention rates of dart tags in their study may be explained by the fact that the posterior position of the dorsal fin in these fish sub- jected the tags to the constant motion of the caudal fin and thus enlarged the wound around the tag in- sertion point (Collins et al., 1994). Compared with other external tags, T-anchor and dart tags appear to have superior retention rates in small red drum (<380 mm SL). Floy dorsal, Monel metal operculum, Monel metal jaw, Petersen oper- culum, Petersen dorsal, streamer dorsal, and Carlin dangler tags all yielded lower retention rates (<50% after 154 DAT, Hein and Shepard, 1980a) than did the T-anchor and dart tags tested in this study. At 66 and 111 DAT, none of the T-anchor tag wounds had healed. Other studies have also shown that tag rotation and movement irritated the inci- sion site, inhibited healing, and left an opening into the body cavity (Smith et al., 1990; Collins et al., 1994). Wallin et al. ( 1997) suggested that long-term retention of T-anchor tags may be reduced because of the open tag wounds; this was not observed in the 423 days of our study. The anchor portion of some of the T-anchor tags protruded through a hole in the abdominal wall that was separate from the tag in- sertion point. This may have been an early stage of tag shedding, although it was never verified that the protruding tags were actually shed. Similar results have also been reported in studies with disk-type internal-anchor tags (Vogelbein and Overstreet, 1987; Mattson et al., 1990). All dart-tag wounds appeared to be healed by 66 DAT. Increases in fish body size during the course of the study caused the streamer portion of some dart tags to become engulfed by the body-wall tissue. This made the tags difficult to detect and impossible to read without dissection. A longer tag streamer could minimize this effect, but it could increase stress on smaller fish during tagging. Simmons and Breuer (1982) also reported tissue growth over tags, which reduced the ability to identify tagged red drum. Tagging did not appear to adversely affect red drum survival. There was no significant difference in in- stantaneous survival estimates between treatments (F=2.2, P=0.26). Mean percent survival of control, T- anchor, and dart-tagged fish was, respectively, 79%, 68%, and 79% at 66 DAT and was 67%, 65%, and 77% at 111 DAT (Table 1). Nearly all mortality oc- curred during the first 40 days of the experiment; 67% of all T-anchor and 83% of all dart-tag mortali- ties occurred during this time period (Fig. 3). These results are consistent with those of other studies in which these tag types were used (Collins et al., 1994; Szedlmayer and Howe, 1995; Wallin et al., 1997). Initial lengths of red drum were not significantly different between treatments (//=3.1, P=0.68, df=5. Table 1 ). Red drum lengths at 66 and 1 1 1 DAT were also not significantly different between treatments (66 DAT: //=3.9, P=0.56, df=5; HI DAT: H=6A, P-Q.21, df=5), suggesting that tagging did not ad- NOTE Winner et a\ : Evaluation of T-anchor and dart tags for use in marking Saaenops ocel/atus 733 100 c O 0. 60 - UO -| — V V .^-v • • ■ Control 2 ^^ V 80 - ■.: ^ 70 - N- ^ ^ 60 - 50 ■ ' Dart 1 Dan 2 00 - 90 ■ ~^.. • • • T-anctior2 80 - V ■ ■ \ _ A 60 - SO - 20 40 60 Days 100 120 Figure 3 Survival of red drum in each treatment group during the first 112 days of the study. versely affect fish growth. It is important that a tag- ging method not affect growth ( Ricker, 1956; Wydoski and Emery, 1983), especially when tagging is being used to verify aging methods (e.g. oxytetracycline injection) or to estimate survival rates in the wild (Green et al., 1985; McFarlane et al., 1990). Growth- rate estimates of red drum in our study (0.41-0.50 mm/day) were lower than those for this size of red drum reported in other studies (Colura and Hysmith, 1975; Hein and Shepard, 1980b), possibly because of the high stocking densities in the net pens. In summary, dart and T-anchor tags are well suited for marking juvenile phase-3 red drum ( 102-173 mm SL) and are usually retained until the fish are large enough to enter the fishery (Florida minimum size limit for red drum is 368 mm SL). Both tag types were easy to apply and did not affect red drum growth or survival. Although tag retention, survival, and growth rates did not differ between tag types, our results did reveal some potential problems associ- ated with the long-term use of T-anchor tags. The wound around the insertion point of the T-anchor tags did not heal during the 423 days of our experiment, and late in the study, some tags showed signs of ex- pulsion (anchor protrusion through abdominal wall). Both tag types could be used to track releases of 734 Fishery Bulletin 97(3), 1999 Table 1 Cumulative percent red drum survival (66 and 111 days after tagging (DAT) and tag retention (66, 111, and 423 DAT) for all treatments. Each treatment was replicated in two separate mesh-net pens, each of which held 50 hatchery-reared red drum. Mean standard lengths (mm) and associated standard errors (in parentheses) are also listed for 0, 66, and 111 DAT Treatment/Pen# % Survival DAT % Tag retention DAT Mean standard length, mm DAT (standard error) 66 111 66 111 423 0 66 111 "Control 1 84 50 — — — 140(2.21) 166 (2.87) 195 (3.65) 2 74 64 — — — 136(2.12) 160(3.13) 191(3.23) T-anchor 3 70 66 98 96 92 138(1.57) 167(2.60) 191(3.05) 6 66 64 98 98 90 141(1.88) 163(2.68) 188(3.91) Dart 4 80 80 98 92 90 140(1.84) 165(2,70) 184(3.71) 5 78 74 100 98 88 138(1.60) 161(2.22) 184(2,28) hatchery-reared phase-3 red drum and thus allow scientists to monitor and accurately determine the effects such releases have on naturally occurring stocks. Acknowledgments We would like thank members of the Stock Enhance- ment and Aquaculture Development team at Port Manatee, Florida, for providing us with the experi- mental fish, aquarium facilities, and expert advice and support that made this study possible. We offer special thanks to Victor Neugebauer for his work maintaining aquarium systems and for feeding and checking fish. We thank Llyn French for her art work and help formatting the manuscript. We also thank Judy Leiby, Jim Quinn, John Ransier, Greg Vermeer, and Julie Wallin for providing helpful comments on the manuscript. This project was funded in part by the Department of the Interior, U.S. Fish and Wild- life Service Federal Aid for sportfish Restoration, Project F-43 to the Florida Department of Environ- mental Protection's Florida Marine Research Insti- tute, and by State of Florida Recreational Saltwater Fishing License funds. Literature cited Arnold, C. R., W. H. Bailey, T. D. Williams, A. Johnson, and J. L. Lasswell. 1977. Laboratory spawning and larval rearing of red drum and southern flounder. Proc. Annu. Conf Southeast, Assoc, Fish Wildl. Agencies 31:437-440, Bumguardner, B. W., R. L. Colura, A. F. Maciorowski, and G. C. Matlock. 1990. Tag retention, survival, and growth of red drum fin- gerlings marked with coded wire tags. Am. Fish. Soc. Symp. 7:286-292. Bumguardner, B. W., R. L. Colura, and G. C. Matlock. 1992. Long-term coded wire tag retention in juvenile Sciaenops ocellatus. Fish. Bull. 90:390-394. Collins, M. R., T. I. J. Smith, and L. D. Heyward. 1994. Effectiveness of six methods of marking juvenile shortnose sturgeons. Prog. Fish-Cult. 56:250-254. Colura, B., and B. Hysmith. 1975. Fingerling production of spotted seatrout, Cynoscion nehulosus. and red drum, Sciaenops ocellatus, in saltwater ponds. Tex. Parks Wildl. Dep. Job Completion Rep. (1975). Elam, L. 1971. Evaluation of fish tagging methods. Tex. Parks Wildl. Coastal Fish. Proj, Rep., p. 39-56. Fox, E., K. Shotton, and C. Ulrich. 1995. SigmaStat users guide, version 2.0, for Windows. .Jandel Scientific, San Rafael, CA, 800 p. Green, A. W., H. R. Osburn, G. C. Matlock, and H. E. Hegen. 1985. Estimated survival rates for immature red drum in northwest Gulf of Mexico bays. Fish. Res. 3:263-277. Gutherz, E. J., B. A. Rohr, and R. V. Minton. 1990. Use of hydroscopic molded nylon dart and internal an- chor tags on red drum. Am. Fish, Soc, Symp, 7:152-160. Hein, S. H., and J. A. Shepard. 1980a. A preliminary tagging study on red drum, Sciaenops ocellatus. in quarter-acre ponds. In Contributions of the Marine Research Laboratory, 1978, p. 33-39. La. Dep. Wildl. Fish. Tech. Bull. 31, (irand Terre Island. Hein, S. H., and J. A. Shepard. 1980b. Growth of juvenile red drum, Sciaenops ocellatus, in quarter-acre ponds. In Contributions of the Marine Research Laboratory, 1978, p. 85. La. Dep. Wildl. Fish. Tech. Bull. 31, Grand Terre Island. Hilborn, R., C. J. Walters, and D. B. Jester Jr. 1990. Value offish marking in fisheries management. Am. Fish. Soc, Symp. 7:5-7. Matlock, G. C, B. T. Hysmith, and R. L. Colura. 1984. Returns of tagged red drum stocked into Matagorda Bay. Texas. Tex. Parks Wildl Manage. Data Ser 63, 6 p. Matlock, G. C, R. J. Kemp Jr., and T. J. Heffernan. 1986. Stocking as a management tool for a red drum fish- ery, a preliminary evaluation. Tex. Parks Wildl, Coastal Fish. Branch Manage. Data Ser. 75, Austin, TX, 27 p. NOTE Winner et al,: Evaluation of T-anchor and dart tags for use in marking Saaenops ocellatus 735 Mattson, M. T., J. R. Waldman, D. J. Dunning, and Q. E. Ross. 1990. Abrasion and protrusion of internal anchor tags in Hudson River striped bass. Am. Fish. Soc. Symp. 7:121- 126. McFarlane, G. A., R. S. Wydoski, and E. D. Prince. 1990. External tags and marks. Am. Fish. Soc. Symp. 7: 9-29. Ricker, W. E. 1956. Uses of marking animals in ecological studies: the marking of fish. Ecology 37:665-670. Roberts, D. E., Jr., B. V. Harpster, and G. E. Henderson. 1978a. Conditioning andinduced spawning of the red drum [Sciacnops ocellatus) under varied conditions of photoperiod and temperature. Proc. World Maricult. Soc. 9:311-332. Roberts, D. E., Jr., L. A. Morey III, G. E. Henderson, and K. R. Halscott. 1978b. The effects of delayed feeding, stocking density, and food density on survival, growth, and production of larval red drum iSciaenops ocellatus). Proc. World Maricult. Soc. 9:333-343. Simmons , E. G., and J. P. Breuer. 1982. Fish tagging on the Texas coast. Tex. Parks Wildl. Dep. Coastal Fish. Proj. Rep., p. 66-107. Smith, T. I. J., S. D. Lamprecht, and J. W. Hall. 1990. Evaluation of tagging techniques for shortnose stur- geon and Atlantic sturgeon. Am. Fish. Soc. Symp. 7:134- 141. Szedlmayer, S. T., and J. C. Howe. 1995. An evaluation of six marking methods for age-0 red drum, Sciaenops ocellatus. Fish. Bull. 93:191-195. Vogelbein, W. K., and R. M. Overstreet. 1987. Histopathology of the internal anchor tag in spot and spotted seatrout. Trans. Am. Fish. Soc. 116:745-756. Wallin, J. E., J. M. Ransier, S. Fox, and R. H. McMichael Jr. 1997. Short-term retention of coded wire and internal an- chor tags in juvenile common snook, Centropomus undecmalis. Fish. Bull. 95:873-878. Weaver, J. E. 1976. Retention of floy tag in red drum, black drum, and sheepshead. Tex. Parks Wildl. Dep. Coastal Fish. Proj. Rep., p. 59-65. Willis, S. A., W. W. Falls, C. W. Dennis, D. E. Roberts, and P. G. Whitchurch. 1995. Assessment of season of release and size at release on recapture rates of hatchery-reared red drum. Am. Fish. Soc. Symp. 15:354-365. Wydoski, R. S. and L. Emery. 1983. Tagging and marking. //; L. Nielsen and D. Johnson (eds.). Fisheries techniques, p. 215-237. Am. Fish. Soc, Bethesda, Maryland. Yokel, B. J. 1966. A contribution to the biology and distribution of the red drum, Sciaejiops ocellata. M.S. thesis, Univ. Miami, Coral Gables, FL, 160 p. 1980. A contribution to the biology and distribution of the red drum,Sc(at'/!ops ocellata. In R. O. Williams, .J. E. Weaver, and F. A. Kalber (eds.). Proceedings of the collo- quium on biology and management of red drum and sea trout, Oct. 19-20, 1978, p. 5. Sponsored by Gulf States Mar. Fish. Comm., Ocean Springs, MS [abstract]. 736 Fishery Bulletin 97(3), 1999 Erratum Erratum: Fishery Bulletin 97(1):80-91 Limburg, Karin E., Michael L. Pace, and Kristin K. Arend ^- Growth, mortality, and recruitment of lar- val Morone spp. in relation to food availabil- ity and temperature in the Hudson River Correction: Karin Limburg has pointed out that the graphs in Figure 2 ( p. 85 ) were mislabeled. Please find below the correct graphs for white perch and striped bass. 0 06 ^ 0 02 c 1 9 -0 02 ^~~-^ /^ adjusted 8 / / Temperature o o white perch [■ striped bass Before During After Before During After Zooplankton bloom status Fishery Bulletin 97(3), 1999 737 Superintendent of Documents Publications Order Form *5178 I I YES, please send me the following publications: Subscriptions to Fishery Bulletin for $43.00 per year ($53.75 foreign) The total cost of my order is $ . Prices include regular domestic postage and handling and are subject to change. (Company or Personal Name) I Please type or print) (Additional address/attention line) (Street address) (City State, ZIP Code) (Daytime phone including area code) (Purchase Order No.) Charge your order ITS EASY! Please Choose Method of Payment: I I Check Payable to the Superintendent of Documents [ I GPO Deposit Account I I VISA or MasterCard Account ■D (Credit card expiration date) (Authorizing Signature) Mail To: Superintendent of Documents P.O. Box 371954, Pittsburgh, PA 15250-7954 To fax your orders (202) 512-2250 Thank you for your order! I I U.S. Department of Commerce Seattle, Washington Volume 97 Number 4 October 1999 Fishery Bulletin The National Marine Fisheries Service (NMFSt does not approve, recommend, or endorse any proprietary product or proprietary material mentioned in this publication. No reference shall be made to NMFS, or to this publication furnished by NMFS. in any advertising or sales promotion which would indicate or imply that NMFS approves, recommends, oi endorses any proprietary product or proprietary material mentioned herein, or which has as its purpose an intent to cause directly or indirectly the adver- tised product to be used or purchased because of this NMFS publication Contents NOV 2 0 1999 y/o~ 738-745 746-757 758-775 776-785 786-801 802-811 812-827 828-841 Anides Aceves-Medina, Gerardo, Enrique A. Gonzalez, and Ricardo J. Saldierna Larval development of Symphurus williamsi (Cynoglossidae: Pleuronectiformes) from the Gulf of California Arreguin-Sanchez, Francisco, and Tony J. Pitcher Catchability estimates and their application to the red grouper (Epinephelus morio) fishery of the Campeche Bank, Mexico Buckel, Jeffrey A., Michael J. Fogarty, and David O. Conover Foraging habits of bluefish, Pomatomus saltatrix, on the U.S. east coast continental shelf Buckel, Jeffrey A., Michael J. Fogarty, and David O. Conover Mutual prey of fish and humans; a comparison of biomass consumed by bluefish, Pomatomus saltatrix, with that harvested by fisheries Friedlander, Alan M., George W. Boehlert, Michael E. Field, Janet E. Mason, James V. Gardner, and Peter Dartnell Sidescan-sonar mapping of benthic trawl marks on the shelf and slope off Eureka, California Gabr, Howaida R., Roger T. Hanlon, Salah G. El-Etreby, and Mahmoud H. Hanafy Reproductive versus somatic tissue growth during the life cycle of the cuttlefish Sepia pharaonis Ehrenberg, 1831 Gillanders, Bronwyn M., Douglas J. Ferrell, and Neil L. Andrew Aging methods for yellowtail kingfish, Seriola iaiandi, and results from age- and size-based growth models Hood, Peter B., and Andrea K. Johnson Age, growth, mortality, and reproduction of vermilion snapper, Rhomboplites aurorubens, from the eastern Gulf of Mexico Fishery Bulletin 97(4), 1999 842-861 Hunt, Joseph J., Wayne T. Stobo, and Frank Almeida Movement of Atlantic cod, Gadus morhua, tagged in the Gulf of Maine area 861 -872 Ibaiiez Aguirre, Ana L., Manuel Gallardo-Cabello, and Xavier Chiappa Carrara Growth analysis of striped mullet, Mugil cephalus, and white mullet, M. curema (Pisces: Mugilidae), in the Gulf of Mexico 873-883 Kaimmer, Stephen M. Direct observations on the hooking behavior of Pacific halibut, HIppoglossus stenolepis 884-890 Lankford, Thomas E., Jr., Timothy E. Targett, and Patrick M. Gaffney Mitochondrial DNA analysis of population structure in the Atlantic croaker, Micropogonlas undulatus (Perciformes: Sciaenidae) 891 -899 Massuti, Enric, Beatriz Morales-Nin, and Joan Moranta Otolith microstructure, age, and growth patterns of dolphin, Coryphaena hippurus, in the western Mediterranean 900-919 Matkin, Craig O., Graeme Ellis, Peter Olesiuk, and Eva Saulitis Association patterns and inferred genealogies of resident killer whales, Orclnus orca, in Prince William Sound, Alaska 920-943 Moser, H. Geoffrey, and Tilman Pommeranz Vertical distribution of eggs and larvae of northern anchovy, Engraulis mordax, and of the larvae of associated fishes at two sites in the Southern California Bight 944-953 Natanson, Lisa J., John G. Casey, Nancy E. Kohler, and Tristram Colket iV Growth of the tiger shark, Galeoceido cuvier, in the western North Atlantic based on tag returns and length frequencies; and a note on the effects of tagging 954-961 Petrik, Rachel, Phillip S. Levin, Gregory W. Stunz, and John Malone Recruitment of Atlantic croaker, Micropogonlas undulatus: Do postsettlement processes disrupt or reinforce initial patterns of settlement? 962-977 Robards, Martin D., John F. Piatt, Arthur B. Kettle, and Alisa A. Abookire Temporal and geographic variation in fish communities of lower Cook Inlet, Alaska 978-986 Simpfendorfer, Colin A. Mortality estimates and demographic analysis for the Australian sharpnose shark, Rhizoprlonodon taylori, from northern Australia 987-998 Smale, Malcolm J., and Andre J. J. Goosen Reproduction and feeding of spotted gully shark, Triakis megalopterus, off the Eastern Cape, South Africa 999-1016 Stoner, Allan W., Allen J. Bejda, John P. Manderson, Beth A. Phelan, Linda L. Stehlik, and Jeffrey P. Pessutti Behavior of winter flounder, Pseudopleuronectes americanus. during the reproductive season: laboratory and field observations on spawning, feeding, and locomotion 1017-1024 Urban R., Jorge, Carlos Alvarez F., Mario Salinas Z., Jeff Jacobsen, Kenneth C. Balcomb III, Armando Jaramillo L., Paloma Ladron de Guevara P., and Anelio Aguayo L. Population size of humpback whale, Megaptera novaeangliae, in waters off the Pacific coast of Mexico 1025-1030 Vecchione, Michael Extraordinary abundance of squid paralarvae in the tropical eastern Pacific Ocean during El Nino of 1987 1031-1042 Watson, William Early life history stages of the whitetip flyingfish, Cheilopogon xenopterus (Gilbert, 1890) (Pisces: Exocoetidae) Notes 1043-1046 Garlich-Miller, Joel L., and Douglas M. Bum Estimating the harvest of Pacific walrus, Odobenus rosmarus divergens, in Alaska 1047-1057 Yang, Mei-Sun The trophic role of Atka mackerel, Pleurogrammus monopterygius, in the Aleutian Islands area 1058-1065 Zeller, Dirk C. Ultrasonic telemetry: its application to coral reef fisheries research 1066 Announcement 1069 Index 1079 Subscnption form 738 Abstract.— The lai-val development of Symphurus williamsi Jordan and Cul- ver, 1895 (Cynoglossidae: Pleuronecti- formes I is described from 69 specimens from the Gulf of California, Me.xico. Di- agnostic characteristics used to identify larvae of this species are 3-5 pigment spots situated on the dorsal region of the body of the base of the pterygio- phore of the dorsal fin and 3-4 pigment spots on the ventral region of the body an the base of the pterygiophore of the anal fin. In early stages, a pigment is present on the base of each anal-fin ray. These pigments also appear at the bases of the dorsal-fin rays in the pos- terior third of ^he dorsal fin. The first three dorsal-fin rays are elongated in lan'ae less than 11.2 mm body length. Larval development of Symphurus williamsi (Cynoglossidae: Pleuronectiformes) from the Gulf of California Gerardo Aceves-Medina Departamento de Plancton y Ecologia Manna Centre Interdisciplinano de Ciencias Marinas Apdo. Postal 592. La Paz, Baia California Sur, Mexico, C P 23000 E-mail address gacevesiffivmredipn ipn mx Enrique A. Gonzalez Departamento de Biologia Marina Universidad Autonoma de Baia California Sur Apdo Postal 19-B La Paz, Baia California Sur, Mexico, C R 23080 Ricardo J. Saldierna Departamento de Plancton y Ecologia Manna Centro Interdisciplinano de Ciencias Mannas Apdo, Postal 592 La Paz, Baia California Sur, Mexico, CP 23000 Manuscript accepted 25 Septemlicr 199fS, Fish. Bull. 97:738-745 ( 1999i. Symphurus is a world-wide genus distributed predominantly in tem- perate and tropical seas and oceans. The genus has 71 recognized spe- cies, 17 of which are found in the American region of the Pacific Ocean (Munroe, 19921. Thirteen species have been found in the Gulf of California: S. atramentatus, S. atri- caudus, S. callopterus, S. chaban- audi, S. eloiigatus, S. fasciolaris, S. gorgonae, S. leei, S. melanuru.s. S. rnelasniatotheca , S. oligomerus, S. prolatinaris, and S. williamsi (Munroe and Mahadeva, 1989; Munroe, 1990; Munroe and Nizin- ski, 1990; Mahadeva and Munroe, 1990; Munroe et al., 1991; Munroe et al., 1995). Specimens of Sym- phurus are usually captured in large quantities by shrimp trawls in some tropical regions, but they are not exploited commercially because of their relatively small size (Perez and Findley, 1985; Van der Heiden, 1985; Van der Heiden et al., 1986). The taxonomy of the eastern Pa- cific species was poorly defined un- til the recent study of Munroe (19921, who concluded that it was possible to identify each of these species by a combination of the dor- sal-, anal-, and caudal-fin ray counts and by the interdigitation patterns of dorsal proximal pterigyophores and neural spines (ID pattern). Symphurus williamsi Jordan and Culver, 1895, can be found from the southern region of the Gulf of Cali- fornia, Mexico, to Panama in shal- low waters with sandy bottoms (Munroe et al., 1995 1. Larvae occur throughout the Gulf of California. Large numbers were found during the summer in collections made south of Islas Tiburon and Angel de la Guarda (Aceves, 1992). Of all species found in the Ameri- can region of the Pacific, only the lar- vae of S. atricaudus (the only species of this genus distributed north as far as California) and S. elongatus have been described (Moser, 1981; Ahl- strom et al., 1984; Matarese et al., 1989; Charter and Moser, 1996). In this study, we offer a description of the larval stages of S. williamsi from preflexion through prejuvenile stages. Obsei'vations made on the lar- val development of S. williamsi are compared with those made previously on S. atricaudus and S. elongatus. Aceves-Medina et al.: Larval development of Symphurus wilhamsi from the Gulf of California 739 Table 1 Meristics for species of Symphurus of the Gulf of California with the same pterygiphore interdigitation pattern ( 1-5-3-2-2) taken from Munroe(1992). S. williamsi S. atricaudus S. chabanaudi S. elongatus S. melasmatotheca S. melanurus S. prolatinaris Dorsal-fin rays 89-95 Anal-fin rays 73-79 Caudal-fin rays 12 Total vertebrae 47-51 94-102 98-109 99-107 90-98 96-104 103-110 77-83 82-92 83-90 74-80 79-87 86-93 12 12 12 11 12 12 50-53 52-57 53-56 49-52 50-54 54-58 Materials and methods Plankton samples were taken during eight oceano- graphic expeditions made in the Gulf of California in April. July-August, and November-December 1984, June and August 1986, July 1991, September 1993, and June 1994. Samples were obtained by us- ing oblique tows with 60-cm diameter bongo nets equipped with nets of 333-|.im and 505-|.ini mesh, as detailed in Kramer et al. (1972). All larvae and juve- niles used in this study were deposited in the larval fish collection of the Plankton and Marine Ecology Department of Centro Interdisciplinario de Ciencias Marinas of the Institute Politecnico Nacional (CICIMAR-IPN). All Symphurus larvae collected with the 505-|im net were separated from the other net contents. Flex- ion, postflexion, and juvenile specimens were cleared, stained (Potthoff, 1984), and identified on the basis of meristic characteristics (Munroe, 1992). Subse- quently, a developmental series of preflexion larvae showing similar pigmentation patterns and meris- tic features was assembled. Counts of the dorsal- and anal-fin elements were sometimes extrapolated from pterygiophore counts, which occasionally appeared earlier in development than did corresponding rays. The following measurements were done on the left side of the body of the specimens: Body depth at anus (BDA); Body length (BL): Head length (HL): Snout-anus length (Sn-A): Body depth (BD): anterior end of snout to end of notochord in young larvae, or to posterior margin of hypural plate in more developed larvae, anterior end of snout to clei- thral margin in small speci- mens, or to posterior margin of the opercle in larger ones. anterior end of snout to poste- rior margin of anus. vertical distance between dor- sal and ventral margins (dor- sal- and anal-fin pterygio- phores excluded) at origin of pelvic fin. vertical distance between dor- sal and ventral margins (dor- sal- and anal-fin pterygio- phores excluded) at origin of anal fin. Eye diameter (ED): horizontal distance between anterior and posterior margins oflefteye(Moser, 1996). The description of larval development was based on 69 specimens (2.2-13.2 mm BL). None of these larvae were in the yolksac stage, indicating that eclo- sion must have taken place at a smaller size. To char- acterize larval development intervals, divisions pro- posed by Ahlstrom et al. ( 1976) were followed. Results Cleared and stained larval and prejuvenile specimens {n = ll, between 7.7-13.2 mm BL) showed the follow- ing pterygiophore patterns: 1-5-3-2-2 (41.2'7f ), 1-5-2- 3-2 (35.2'7f), 1-5-2-2-2 (11.8^f), 1-4-2-2-2 (5.9'7f). and 1-5-4-2-2 (5.9%). There are seven species in the area with the same main pterygiophore pattern (Table 1). Using these 17 specimens and the remain- ing 52 larvae (Table 2), we counted 9 precaudal ver- tebrae, 38-42 caudal vertebrae (mode=40), 89-97 dorsal rays (mode=92 and 94), 71-80 anal rays (mode=75), and 12 caudal-fin rays. The combination of these characteristics indicates that the species is S. williamsi. Description of the larvae Diagnosis Symphurus williamsi larvae typically have a pigmentation pattern of 5 spots along the middorsal line at the bases of the dorsal-fin pterygiophores and 2 to 3 similar spots located on the bases of the dorsal-fin rays. There are also 3 to 4 740 Fishery Bulletin 97(4), 1999 Table 2 Symphurus williamsi lai-vae and juvenile measurements (mm): body length (BL). snou t-anus length (Sn-A), head length (HL), | eye diameter (ED) body depth (BDl body depth at anal fin (BDA), and individual counts dorsal rays(DR), anal rays (AR), caudal rays(CR), caudal vertebrae (CV), ar d total vertebrae (TV) Start of flexion stage ( : start of postflexion stage ( - - - ); start of juvenile stage ( ). BL Sn-A HL ED BD BDA DR AR CR CV TV 2.2 1.1 0.6 0.2 0.5 0.3 22 31 2.4 1.1 0.6 0.2 0.5 0.3 37 46 -^ 2.6 11 0.6 0.2 0.7 0.3 32 41 2.8 1.2 0.7 0.2 0.9 0.6 38 47 2.8 1.2 0.7 0.2 0.8 0.4 34 43 2.9 1.3 0.7 0.2 0.9 0.7 40 49 3.1 1.5 0.8 0.2 1.1 0.7 38 47 3.2 1.4 0.9 0.2 1.2 0.8 39 48 3.3 1.7 1.1 0.2 1.4 1.1 38 47 3.5 1.5 0.9 0.2 1.1 0.7 40 49 3.7 1.5 1.1 0.2 1.3 0.9 40 49 3.8 1.8 1.1 0.2 1.2 0.9 38 47 4.0 1.9 1.2 0.2 1.3 0.9 38 47 4.1 1.9 1.3 0.2 1.4 1.1 40 49 4.3 1.9 1.1 0.3 1.6 1.2 40 49 4.3 1.9 1.1 0.3 1.6 1.1 40 49 4.5 2.1 1.2 0.2 1.8 1.1 41 50 4.6 2.0 1.3 0.2 1.5 1.2 40 49 4.8 1.8 1.1 0.2 1.5 0.9 39 48 5.0 2.0 1.2 0.3 1.7 1.4 39 48 5.0 1.9 1.2 0.2 1.7 1.1 40 49 5.0 2.1 1.3 0.3 1.6 1.2 38 47 5.3 2.1 1.4 0.3 1.7 1.2 40 49 5.6 2.0 1.4 0.3 1.5 1.1 41 50 5.8 2.3 1.5 0.3 1.9 1.4 40 49 5.9 2.3 1.4 0.3 1.9 1.2 40 49 6.1 2.4 1.6 0.3 1.9 1.5 40 49 6.1 2.2 1.7 0.3 1.9 1.4 80 40 49 6.3 2.4 1.5 0.3 1.8 1.4 40 49 6.6 2.2 1.6 0,3 2.1 1.5 38 47 6.8 2.5 1.7 0.3 2.3 1.5 92 79 41 50 7.0 2.6 1.6 0.3 1.9 1.5 92 77 41 50 7.2 2.7 1.8 0.3 2.2 1.7 92 75 40 49 7.3 2.7 1.7 0.3 1.9 1.5 95 79 40 49 7.4 2.5 1.7 0.3 2.3 1.5 94 75 41 50 7.7 2.6 1.7 0.3 2.0 1.7 92 75 12 40 49 7.8 2.9 1.9 0.3 2.0 1.6 94 76 12 38 47 8.0 2.8 2.0 0.3 2.2 1.7 93 77 41 50 8.2 2.8 1.9 0.3 2.3 1.9 96 77 12 40 49 8.3 3.1 2.0 0.3 2.4 1.9 95 76 12 41 50 8.3 2.9 1.9 0.3 2.4 1.8 96 79 12 40 49 8.3 2.9 1.9 0.3 2.3 1.9 93 76 12 40 49 8.4 3.0 1.9 0.3 2.1 1.5 94 76 12 39 48 8.7 2.9 2.2 0.3 2.5 2.1 91 74 12 40 49 9.3 3.0 2.1 0.3 2.0 1.6 94 75 12 38 47 9.5 3.3 2.2 0.3 2.3 2.0 94 77 12 40 49 9.6 3.4 2.2 0.3 2.6 2.1 91 75 12 38 47 9.8 3.1 2.0 0.3 2.3 2.0 92 76 12 40 49 9.9 3.2 2.1 0.3 2.2 2.0 93 71 12 39 48 9.9 3.2 2.1 0.3 2.3 2.1 94 78 12 41 50 10.0 3.0 2.2 0.3 2.5 2.2 91 75 12 40 49 10.2 2.8 2.2 0.3 2.4 2.2 92 75 12 41 50 continued Aceves-Medina et al.: Larval development of Symphurus williamst from tfie Gulf of California 741 Table 2 (continued) BL Sn-A HL ED BD BDA DR AR CR CV TV 10.4 3.1 2.2 0.4 2.5 2.3 92 76 12 41 50 10.6 3.4 2.3 0.4 2.3 2.1 93 74 12 38 47 10.6 3.3 2.3 0.3 2.2 2.2 91 75 12 39 48 10.7 3,5 2.2 0.3 2.6 2.3 94 76 12 10.8 3.3 2.5 0.4 2.6 2.4 91 73 12 39 48 11.1 3.4 2.4 0.3 2.5 2.2 94 77 12 39 48 11.1 3.6 2.5 0,3 2.7 2.7 90 75 12 38 47 11.1 3.7 2.6 0,3 2.4 2.4 97 80 12 39 48 11.1 3.7 2.5 0,3 2.4 2.4 93 76 12 42 51 11.3 3.5 2.5 0.3 3.0 2.3 92 77 12 39 48 11.4 3.3 2.3 0.3 2.5 2.3 94 79 12 40 49 11.7 3,4 3.2 0.4 S.'-i 3.3 93 73 12 11.9 3.8 2,8 0,3 2.5 2.3 92 12 40 49 12,7 3.2 2.5 0,4 2.3 2.7 91 75 12 38 47 12.9 3.9 3.0 0,3 2.4 2.6 89 74 12 39 48 13.1 3.9 3.0 0,3 3.1 2.4 94 74 12 13.2 3.4 2.9 0,3 2.2 2.9 92 75 12 39 48 spots along the midventral line at the bases of the anal-fin pterygiophores, and 1 to 2 similar spots lo- cated on the anal-fin ray bases. At time of formation of the anal-fin rays, a melanophore appears at the base of each fin ray. A similar pigment is also found at the base of each dorsal- fin ray in the posterior third of the body in larvae less than 11.2 mm BL. The first 3 dorsal rays are elongated in larvae less than 11.2 mm BL, with the second and third of these longer than the first. Morphological features The body is compressed and ribbonlike and has a deep head. The gut extends be- yond the ventral profile. The intestine is noticeably coiled and its terminal end is directed towards the right side of the body. Larvae attain the appearance of juveniles when the protruding gut is incorporated within the ventral body profile at approximately 12 mm BL. At metamorphosis, a fleshy rostral beak is formed anterior to the dorsal-fin origin. Migration of the right eye starts at 6.1 mm BL and eye migration is completed in larvae of 11.1 mm BL. The migrating eye crosses anterior to the dorsal-fin origin, between the rostral beak and the interorbital region. Notochord flexion begins at 5.0 mm BL and is completed near 8.0 mm BL. The juvenile stage begms at about 11.9 mm BL (Table 2). Growth is not isometric throughout larval devel- opment (Table 3). Head length, snout-anus length, and body depth decrease with respect to body length from preflexion to juvenile stages. Body depth at the anal-fin origin remains constant throughout devel- "opment. Eye diameter in relation to head length de- creases from preflexion to juvenile stages. Table 3 Symphurus wi Itarnsi larvae andj uvenile bodv proportions (in percent of BL); body len^h (BL , head 1 ength (HL), snout-anus len gth (Sn-A), body depth (BD). body depth at 1 anal fin (BDA) , and eye diameter (ED). Standard Body proportions Average deviation Interval HL:BL Preflexion 27 2.8 23-33 Flexion 25 1.3 23-28 Postflexion 23 1.5 20-27 Juvenile 22 1.4 20-24 Sn-A:BL Preflexion 45 3.3 38-52 Flexion 38 2.2 33-42 Postflexion 33 2.4 28-37 Juvenile 29 2.6 25-32 BD:BL Preflexion 33 5.2 21-42 Flexion 31 2.5 26-34 Postflexion 25 2.6 21-30 Juvenile 20 2.5 17-24 BDA:BL Preflexion 22 5.5 12-33 Flexion 23 2.0 20-28 Postflexion 22 2.0 17-28 Juvenile 20 1.3 18-23 ED:HL Preflexion 24 6.2 15-33 Flexion 19 2.5 15-25 Postflexion 14 1.8 12-18 Juvenile 11 2.3 10-12 742 Fishery Bulletin 97(4), 1999 Formation of scales begins at 7.7 mm BL over the cephalic region, and by about 10.3 mm BL, scales are evident over the entire body. Pigmentation Early preflexion larvae typically have three melanophores along the middorsal line and two or three on the dorsal-fm fold. Three other melano- phores are present along the midventral line, and one or two similar spots are found on the anal-fin fold. A illelanophore is located over the coiled intestine in anterior and posterior positions. The ventral surface of the coiled intestine shows numerous smaller and dispersed melanophores. Two other melanophores ap- pear in the anterior region: one on the cleithrum near the base of the pectoral bud; the other on the ventral margin of the preopercle. The dorsal surface of the swimbladder is also pigmented (Fig. lA). With fmfold disappearance (late preflexion larvae), the characteristic pigmentation pattern is estab- lished: five spots located along the middorsal line at the bases of the dorsal-fm pterygiophores and two or three spots on the bases of dorsal-fin rays. Three or four spots appear along the midventral line at the bases of the anal-fm pterygiophores, and one or two spots are present on the bases of anal-fin rays. When formation of the dorsal- and anal-fin rays is complete, a pigment spot appears on the base of each anal ray. Similar pigments are found on the bases of the fin rays in the posterior third of the dorsal fin. On the head, an inner melanophore appears over the forebrain and another over the otic region. Finally, a melanophore is present on the distal margin of the peduncle of the pectoral fin (Fig. IB). During the flexion stage, the pigmentation pattern of preflexion lai^vae remains, but the spots on the bases of anal- and dorsal-fin rays sometimes disap- pear. The pigment over the otic region, that on the opercle margin, and the melanophore on the poste- rior part of the intestine also disappear. In this stage there are additional pigments that appear. A series of dots are found over the ventral margin of the head, and a melanophore appears over the inferior portion of the base of the ventral fin. Flexion larvae are simi- lar to those of postflexion larvae (Fig. 2A). At the postflexion stage, three melanophores appears on the bases of the caudal-fin rays, and at 11.2 mm BL, the cephalic and opercular pigments may disappear (Fig. 2A). In early juveniles, four spots ap- pear on the midline of the body. Their positions with respect to the melano- phores already present on the dorsal margin are one between second and third spots, one between third and fourth, one directly ventral to the fourth spot, and one posterior to the fourth spot. There is now pigment on the base of every dorsal-fin ray throughout the entire fin. A series of small spots also develop on the rays on the anterior part of the anal fin (Fig. 2B). Fin formation Unpaired fin development Incipient dorsal-fin rays are present in the an- terior dorsal finfold at 2.8 mm BL. At 4.0 mm BL, about one half of the dor- sal and anal rays are formed in a pos- terior direction. Complete formation of these fins occurs at about 6.8 mm BL. The 3 elongated dorsal-fm rays, present since this fin originated, dis- appear at 11.4 mm BL. Figure 1 Symphurus williamsi: (A) 2.2-mm preflexion larva; (B) 6.3-mm flexion larva. Paired fin development Pectoral-fin buds are present in the smallest lar- I Aceves-Medina et al.: Larval development of Symphurus williamsi from thie Gulf of California 743 vae we have seen (2.2 mm BL). This fin is reabsorbed at the postflexion stage (10.3 mm BL). Pelvic fins are de- veloped at 5.7 mm BL. Only in early juveniles is a blind- side pelvic fin located on the midventral line and the ocu- lar pelvic fin is in a more dor- sal position. Discussion Larvae of S. williamsi are easily distinguishable from those of S. atricaudus by dif- ferences in pigmentation pat- terns. During early preflex- ion stage, S. williamsi has two or three pigments over the dorsal and midventral lines, and the pigments over the finfold are in the middle. S. atricaudus has a series of small melanophores along the midventral line. The pig- ments over the finfold are in blotches and are on the ven- tral and dorsal margins. There are also several pig- ments over the ventral margin of the head and the gut that S. williamsi does not have. At the flexion stage, S. williarrjsi has the spot pat- tern along the dorsal and ventral profile. In contrast, S. atricaudus has a series of melanophores along the dorsal- and anal-fin margins and a series of pigments along the midbody line that never develop in S. williamsi. In S. williamsi. three elongated rays are charac- teristic. In S. atricaudus, five rays appear first in the dorsal fin but never become elongated. The in- testine in S. williamsi projects only ventrally, whereas in S. atricaudus the projection is ventro- posterioriad. Larvae of S. williamsi can be distinguished from those of S. elongatus by the pigmentation patterns. In the preflexion stage, S. elongatus does not have spots at the bases of the dorsal-fin rays, and these larvae do not have pigments at the bases of each of the anal- and dorsal-fin rays. Later in their develop- ment, the pigmentation of S. elongatus consists of a few distal pigments on the anal fin, twelve dashes along the lateral midline, and a small blotch on the caudal-fin rays, not observed in S. williamsi. In the early transformation stage, S. elongatus has a series of Figure 2 Symphurus wUliamsr. ( Al 10.0-mm postflexion larva; (B) 11.9-mm transformation larva. melanophores over the anterior part of the head, which never appear in S. williamsi. Like S. atricaudus, S. elongatus never develops elongated rays. Yevseyenko (1990) described Symphurus sp. lar- vae. Its meristic characters have led us to believe it is S. callopterus. principally because of its ID pat- tern. These larvae are different from those of S. williamsi in pigmentation pattern because on the caudal section of the body an accumulation of pig- ment in the form of two oblique bands and two belts are found. These pigments are approximately equi- distant from one other and never are found in S. willianisi. Instead of three elongated dorsal-fin rays, like S. willia/nsi, Symphurus sp. have the first 7 dorsal-fin rays elongated from 7.5 to 25.4 mm BL (Yevseyenco, 1990). The presence of elongated fin rays in pleuronectiform larvae has phylogenetic interpreta- tions only in bothids and appears to be apomorphic within the order and this family (Hensley and Ahlstrom, 1984). However the use of this character- istic for phylogenetic interpretations in cynoglossids does not allow proper comparisons because most lar- vae remain unknown. Adults of S. williamsi (7.5 to 11.6 cm BL) are smaller than those of S. atricauda (13 to 18 cm BL) 744 Fishery Bulletin 97(4), 1999 and S. elongatus (13.5-15 cm BL) (Munroe et al., 1995), and this size difference is observed in larvae also. All stages of development in S. williamsi occur at smaller sizes (flexion between 5.0 and 8.0 mm and transformation >11.9 mm) than in other species. The flexion stage in S. atricaudus occurs between 8.6 and 10.8 mm BL (Ahlstrom et al., 1984; Matarese et al., 1989), and the transformation stage between 19 and 24.2 mm BL (Kramer, 1991 ). The flexion stage in S. elongatus is >6.8 mm and <18.9 mm BL and trans- formation takes place from >18.9 mm to <28.3 mm BL (Charter and Moser, 1996). In Symphurus sp. transformation occurs between 21 and 38 mm BL (Yevseyenko, 1990). Usually the size range at metamorphosis that is plesiomorphic for the order is ca. 10 to 25 mm BL because most pleuronectiforms metamorphose in this range (Hensley and Ahlstrom, 1984). Metamorphosis in S. williamsi occurs at ca. 11.9 mm BL; however, as described by Hensley and Ahlstrom ( 1984), size at metamorphosis is an impor- tant characteristic for lai-val identification, but its use for inferring phylogenetic relations in most in- stances is premature. Symphurus williamsi has a larger head at all de- velopmental stages, compared with those of S. atricaudus and S. elongatus. The distance from snout to anus in S. williamsi is relatively longer than it is in S. atricaudus but is similar to that of S. elongatus. In the juvenile stage, this distance is longer in S. williamsi than it is in S. elongatus. Larvae of Symphurus sp. as described by Yevseyenko (1990) have an abdominal cavity extending downward. This projection has the form of a greatly elongated sac, and the lower end terminates in a wormlike process that S. williamsi does not have. The loop of the in- testine in Symphurus sp. is almost one half of the body length. The body depth at the preflexion stage is greater in S. williamsi compared with that in S. atricaudus but is similar to that of larvae of S. elongatus. In postflexion and prejuvenile stages, this proportion is smaller in S. williamsi than in both of these other species. Finally, body depth at the anus is shallower in S. williamsi than it is in S. elongatus at all stages in S. atricaudus during flexion and juvenile stages. Acknowledgments This study was funded by the Direccion de Estudios Profesionales - I.P.N, through the following research projects: Planton del Noroeste de Mexico (Clave DEPI 86804); Investigaciones Ecologicas del Planton del Noroeste de Mexico (DEPI 868043); Caracterizacion de la Zona de Transicion Templado-Tropical del Pacifico Mexicano con base en las Comunidades Plantonicas (DEPI 874264); and Bionomia Plantonica de la Parte Central del Golfo de California (Clave DEPI 903388). We especially thank our colleague Andres Levy who assisted in critically revising the manuscript and translating it into English, Biol. Raymundo Avendaho Ibarra who made the drawings, G. Moser (Southwest Fisheries Science Center) who provided helpful comments that improved the manu- script, and Elhs Glazier, CIBNOR, who edited the English-language text. Literature cited Aceves, M. G. 1992. Analisis espacio temporal de la distribucibn y abundancia de larvas de Pleuronectiformes en el Golfo de California, periodo 1984-1986. M.S. thesis. Centre Interdisciplinario de Ciencias Marinas-IPN. La Paz. B.C.S. Mexico. 62 p. Ahlstrom, E. H. 1976. Maintenance of quality in fish eggs and larvae col- lected during plankton hauls. In H. F. Steedman (ed), Zooplankton fixation and preservation, p. 313-3 18. Unesco Press, Paris. Ahlstrom, E. H., K. Amaoka, D. A. Hensley, H. G. Moser, and B. Y. Sumida. 1984. Pleuronectiformes: development, /fi H. G. Moser, W. J. Richards, D. M. Cohen, M. P. Fahay, A. W. Kendall, and S. L. Richardson (eds. I, Ontogeny and systematic of fishes, p. 640-670. Spec. Publ. 1. Am. Soc. Ichthvol. Herpetol. Charter, S. R., and H. G. Moser. 1996. Cynoglossidae: tonguefishes. In H. G. Moser led I, The early stages of fishes in the California current region. CALCOFI ATLAS 33, p 1408-1413. Allen Press, Inc., Lawrence, KS. Hensley, D. A., and E. H, Ahlstrom. 1984. Pleuronectiformes: relationships. In H. G. Moser, W. J. Richards, D. M. Cohen, M. P Fahay, A. W. Kendall, and S. L. Richardson I eds.). Ontogeny and systematic of fishes, p. 640-670. Spec. Pub. 1. Am. Soc. Ichthyol. Herpetol. Kramer, D., M. J. Klain, E. G. Stevens, J. R. Thraikill, and J. R. Zweifel. 1972. Collecting and processing data from fish eggs and larvae in the California Current Region. LI.S. Dep. Conimer. NOAA Tech. Rep. NMFS Circ. 370. 38 p. Kramer, S. H. 1991. The shallow-water flatfishes of San Diego County. Calif Coop. Oceanic Fish. Invest. (CALCOFI) Rep. 32:128- 142. Mahadeva, M. N., and T. A. Munroe. 1990. Three new species of symphurine tonguefishes from tropical and warm temperate waters of the eastern Pacific iSymphiirus: Cynoglossidae: Pleuronectiformes i. Proc. Biol, Soc. Wash. 103U i:931-9.-54. Matarese, A. C, A. W. Kendall, Jr., D. M. Blood, and B. M. Vinter. 1989. Laboratory guide to early life history stages of North- Aceves-Medina et al.: Larval development of Symphurus williamsi from tfie Gulf of California 745 east Pacific fishes. U.S. Dep. Commer., NOAA Tech. Rep. NMFS 80, 652 p. Moser, H. G. 1981. Morphological and functional aspects of marine fish larvae. In R. Lasker (ed). Marine fish larvae: morphol- og>'. ecology, and relation to fisheries, p. 89-1.31. Univ. Washington Press, Seattle, WA. 1996. The early stages of fishes in the California current region. CALCOFI atlas 33. Allen Press, Inc., Lawrence, KS, 1505 p. Munroe, T. A. 1990. Symphurus melanurus Clark, 1936. a senior synonym for the eastern Pacific tonguefishes, S. seychellensis Chaba- naud. 1955 and S. sechurae Hildebrand, 1946. Copeia 1990(11:229-232. 1992. Interdigitation pattern of dorsal-fin pterygiophores and neural spines, an important diagnostic character for Symphurinae tonguefishes (S,vmp/!ur;is:Cynoglossidae: Pleuronectiformes). Bull. Mar. Sci. 50(3 1:357-403. Munroe, T. A., F. Krupp, and M. Schneider. 1995. Cvnoglossidae. In W. Fischer, F. Krupp, W. Schneider, K. Carpenter, C. Summer, and V. Niem (eds.). FAO species identification sheets for fisheries purposes: eastern cen- tral Pacific ( fishing area 77 1, p. 1039-1059. Food and Agri- culture of the United Nations, Rome, Italy; Senckenberg Re- search Institute, Frankfurt, Germany. Munroe, T. A., and M. N. Mahadeva. 1989. Symphurus callopterus (Cynoglossidae, Pleuronecti- formes), a new deepwater tonguefish from the eastern Pacific. Proc. Biol. Soc. Wash. 102(2i:458-467. Munroe, T. A., and M. S. Nizinski. 1990. Symphurus melasmatotheca and S. undecimplerus (Cynoglossidae, Pleuronectiformes), two new eastern Pa- cific tonguefishes with eleven caudal-fin rays. Copeia 1990(4):985-996. Munroe, T. A., M. S. Nizinski, and M. N. Mahadeva. 1991. Symphurus prolatinaris. a new species of shallow- water tonguefish (Pleuronectiformes: Cynoglossidae) from the eastern Pacific. Proc. Biol. Soc. Wash. 104(3):448-458. Perez, M. J., and L. T. Findley. 1985. Evaluacion de la ictiofauna acompanante del camaron comercial capturado en las costas de Sonora y norte de Sin- aloa. In A. A. Yanes ( ed. ), Recursos pesqueros potenciales de Mexico: la pesca acompafiante del camaron, p. 201- 251. Programa Universitario de Alimentacion. Instituto de Ciencias del Mar y Limnologia, Instituto Nacional de la Pesca., UNAM, Mexico, D. F. Potthoff, T. 1984. Clearing and staining techniques. In H. G. Moser. W. J. Richards, D. M. Cohen, M. P Fahay, A. W. Kendall and S. L. Richardson (eds.). Ontogeny and systematic of fishes, p. 3.5-37. Spec. Publ. 1. Am. Soc. Ichthyol. Herpetol. Van der Heiden, A. M. 1985. Taxonomia biologia y evaluacion de la ictiofauna de- mersal del Golfo de California. In A. A. Yanez (ed.), Recursos pesqueros potenciales de Mexico: la pesca acompaiiante del camaron, p. 149-200. Programa Universitario de Alimentacion, Instituto de Ciencias del Mar y limnologia, Instituto Nacional de la Pesca., UNAM, Mexico, D.F. Van der Heiden, A. M., H. Plasencia, and S. Mussot 1986. Aportaciones al conocimiento de la ictiofauna dem- ersal del Golfo de California. In Memorias de la I Reunion Interc. Acad. Inves. del Mar de Cortes, Hermosillo, Mexico, p. 327-339. Yevseyenko, S. A. 1990. Unusual larvae of the marine tonguefish, Symphurus sp. (Cynoglossidae), from central waters of the eastern Pacific. Voprosy Ikhtiologii 30(41:682-686. 746 Abstract.— We applied a length-based additive catchability model that ac- counts for several sources of variation, namely time (i.e. years, months!, den- sity-dependence effects, and different fishing fleets. The model is based on 1) a transition matrix and 2i population length-structured data expressed as catch per unit of effort. Other sources of variation were estimated as anoma- lies from the average pattern and in- corporated as additions to the slope of the catchability-at-length equation. The catchability model was applied to the red grouper \Epinephelus morio) fishery of the north continental shelf of Yucatan, a demersal fish resource ex- ploited by three different fleets. A sig- moidal shape catchability-at-length function was fitted on the basis of grou- per population biology and behavior, particularly reproductive aggregation. Catchability of immature fish was con- stant but increased with size for adult fish, especially during the reproductive season. Time- and density-dependent catchability responded to reproductive behavior and the allocation of fishing effort. When differences between fleets were incorporated, catchability differ- ences emerged. The catchability model has the ability to identify the main properties of the fish resource and fish- ery that affect the relation between fishing effort and population abun- dance; it may therefore be helpful as an alternative stock assessment tool. Catchability estimates and their application to the red grouper (Epinephelus morio) fishery of the Campeche Bank, Mexico Francisco Arreguin-Sanchez Centre Interdisciplinano de Ciencias Mannas del IPN, CICIMAR Apartado Postal 592, 23000, La Paz, Baia California Sur, Mexico E-mail address, farreguiffflvmredipn ipn mx Tony J. Pitcher Fisheries Centre University of British Columbia Vancouver, V6T 1Z4, Canada Manuscript accepted 19 November 1998. Fish. Bull. 97:746-757 (1999). Catchability has been considered in fisheries science as a parameter in the catch equation that relates fish- ing effort to population abundance: C = qsEN, (1) where C = catch in numbers; q - catchability; s = probability of gear selection; E = fishing effort; and N = the stock size in numbers. From this relation it is easy to un- derstand the key role of g. Because s and E are controlled by man, changes in population abundance will be reflected in catch through q. Although classically considered a constant, q represents different sources of variation affecting stock size. Most of the existing catcha- bility models deal with only one source of variation related to school- ing of fish (e.g. MacCall, 1976; Csirke, 1988, 1989), although a few of them also consider environmen- tal factors (Table 1), Investigations on catchability have developed mainly in two direc- tions (Arreguin-Sanchez, 1996): 1) those related to measuring and in- creasing gear efficiency, and 2 ) those that use catchability as a parameter to relate fishing effort to fishing mortality and population abun- dance for stock assessment and management purposes. The aim of this contribution is to apply a catchability model that accounts for several sources of variation related to fishing and to population pro- cesses, as proposed by Arreguin- Sanchez (1996). The red grouper (Epinephelus morio) stock over the Campeche Bank is one of the largest in the world, distributed on a continental shelf of more than 100,000 km^. As with many serranids. the red grou- per stock over the Campeche Bank, Gulf of Mexico, aggregates for re- production, a fact that is well docu- mented in the literature (Arreguin- Sanchez et al., 1996). While they aggregate, these fish are highly vul- nerable to fishing; therefore this fish behavior is a key aspect for management of the resource. Three fleets target this fish species: two from Mexico (an artisanal and a midsize fleet); and a third from Cuba. These fleets do not com- pletely overlap in respect to their fishing grounds, and their catch structure, efficiency, and the fish- ing mortality that they cause to the stock are different. Our contribution is aimed to estimate catchability for the red grouper fishery. Arreguin-Sanchez and Pitcher: Catchability estimates for the Epinephelus mono fishery 747 Table 1 Catchability models in literature accounting for more than one source of variation. Catchability model Comments Reference <7, = 9o exp-" r, = a, (t) T.,= a, E(t) i=time, E= fishing effort Quinn(1987) q = 0.085 W-o 384 q = 0.287 AT-" 2'8 Haddock, George's Bank A' = population abundance Area = population habitat Grecco and Overholtz ( 1990) q = 0.095 AA-o "8" q = 0.003 Area-'^^*'' Cod gill net, Atlantic S=Stock biomass Area = population habitat Rose and Legget(1991) T/q. Peruvian anchoveta. q^ = base fraction of the stock size per unit of effort (when stock size is low); B = biomass; T = temperature; q., and q^ are constants Hilborn and Walters ( 1992) qi 1 .t.E. F) = /I size) f = length, t = time. E = fishing effort; f = fieets Arreguin-Sanchez (1996) Materials and methods We assume that catchability 1 ) is length dependent; 2) depends on fish behavior; 3) and is density depen- dent. For fishing effort, we also assume that 1 ) these assumptions can be valid for individual fishing fleets, such that we can explicitly consider the individual contribution of the several fleets participating in the fishing process and 2) that units of effort have static properties, that is to say, a consistent fishing power over time. where G = the effect of growth in absence of mor- tality; and S = theeffectofmortality (survival) and also of selection of the sampling gear. Matrix G( i ,k) can be easily estimated by following the criteria defined by Shepherd (1987) to assign growth probabilities to each length class. The ele- ments of S(/f)can be defined in terms of mortality as S(k} : -Zi/ri( -lM+ol*./ls(*l£'(n| e = e ' ', Length-dependent catchability The catchability coefficient must be estimated for each length class in a given time, qi i ,t). A convenient form to represent the transformation of one length- frequency distribution (representing the structure of the stock) into another is by means of a transition matrix (Shepherd, 1987; Caswell, 1988), so that N(i.f + l) = A{i,k)N{i,t), (2) where k and ' = successive length intervals; N( I ,t) = is the stock size in numbers at time t; and A = the transition matrix that depends on growth and mortality. Although these processes occur concurrently. Shep- herd ( 1987) assumed that they can be separated as the product of two terms: A{i,k) = Gii,k)S{k), where S(k) represents the values of the elements in the main diagonal of the survivorship matrix (other elements are zero; see Caswell, 1988); Z{k)t is the in- stantaneous rate of total mortality for the /c"^ length group at time t; M is the instantaneous rate of natural mortality (assumed constant); s(k) is the probability of gear selection for length k\ and E(t) is the fishing effort at time / ( which is assumed nontargeted to specific sizes within a range of sizes captured by the gear). Then fish- ing mortality is given as Fik. t) = q(k, t) sik)E(t). Equation 2 can therefore be represented as N(i ,t + h = Y^G(i ,k)e-^"^'"'--"""""^N(k,t). (3) If A'^( (t-i-1), NikJ) and Gi i ,k) are known, as well as M,s(k). and E(t), then g(/e,/) can be estimated. Equa- tion 2 can then be solved iteratively for qik.t) by us- ing an algorithm to minimize differences between observed and calculated values o^ Ni i . t + 1). Once q(k.t) values are obtained, the catchability pattern with length can be observed and a tendency 748 Fishery Bulletin 97(4), 1999 quantified. Although a catchabiHty-at-length model may be empirically determined, its interpretation must be made by considering the population biology and behavior of the fish. On an annual basis, the model is represented by the function qiOaPin-f (4) whose slope ^( ' ) represents the rate of change of catchability with length for a given year. If several years are considered for a stable period of the fish- er}', then a general function can be estimated where the slope of the length-based catchability pattern in Equation 4 can be expressed by /3( ',•), where the symbol • represents the average stock level over a range of years in which it can be reasonably assumed to be stable. Time-dependent catchability GuUand ( 1983 ) expanded the scope of the catchability coefficient to a structured population as expressed in the following equation: C{l,At) = q{l,At)s(nE{At)N{i,At)- (5) This represents the capture of fish of various sizes ( ' ) during a time period Dt. Taking the mean stock size ( A'^ ), and rewriting Equation 5 in terms of catch per unit of fishing effort, U, we obtain ith lHi.') = q(i,»}N{i), c/(/,•) = C(^An/.s(/)£(An and then Ln[L^( I ,t)/U{ ',•)] = Ln[(7( i,t)N(i )/q(l.»)N{ i )] = Ln[q{i .t)/q(i .•)]. (6) This can be interpreted as the departure or anomaly in catchability at time t with respect to the average for a given length class. Arreguin-Sanchez (1996) suggested that this ratio is a linear function of size, represented by the midlength class as where Ln[g( ',n/(7( /,•)] = am + /J(n( , (7) p(t) = Ln[q(i + lJ)/q(i,t}]-Ln\qn + 1,« )/(/(', •)]. The intercept cdt) in Equation 7 can be interpreted as the relative vulnerability of small fish, and also as an index of the relative abundance of recruits. The slope (Xt) expresses the rate of change of catchability- at-length with respect to the average, or in other words, the departure (anomaly) from the average catchability pattern with length. Once /^( ' ,• ) is estimated from Equation 4, the value of j3(^) can be added to pi i .*) to obtain the correspond- ing catchability values for a given time period. Amount of fishing and density-dependent catchability The relation between population size and catchability has been most studied in pelagic clupeoid fish (Murphy, 1966; MacCall, 1976; Csirke, 1988, 1989; Pitcher, 1996), where catchability is inversely related to stock size. Thus a density-dependent effect can be approached by considering the population as a whole, and the function qw(t) = f{N(t)} (8) can be fitted, where qwtt) is the weighted value of g at time t, with U at time ^ as a weighting factor. An analogous approach is proposed for each length class. For a stable period of time, we can assume that the amount of fishing, or the total fishing effort E, is one of the main sources of variation of the population density and the structure of the exploited stock. A change in both of these population characteristics will be reflected in the pit) coefficient. With Equa- tions 1 and 8, it is possible to relate changes in qU ,t) as a function of coefficient fi(t): pu)aP(E)i, (9) where a constant natural mortality rate is assumed for all length classes over time. Because pit) represents the departure of the catchability pattern with length at time t. Equation 9 measures the effect of the amount of fishing on the anomaly, and it is represented by the slope /j(£). The magnitude and sign of piE) will provide information about changes in stock density. A negative sign means that for a low level of fishing, q will increase with fish length. Catchability differences between fleets Following the same rationale as that in Equation 7, differences in q between fleets as a function of length can be estimated. If c/ and h represent two different fishing fleets, Equation 6 can now be expressed as Ln[lHg,iJ)/Uih,iJ)] = Ln[q{g,i.t)/q{h.i,t)\ (10 I and Aiieguin-Sanchez and Pitcher: Catchability estimates for the Epinephelus mono fishery 749 Ln[q{gJ,f)/q{hj,t}] = a(f) + [i(f)(, (11) where /" indexes fishing fleets. In this case, the intercept a(/") is the relative dif- ference in catchabiUty between fleets for the small- est fish. The slope Pif) reflects the difference in fish- ing performance measured as the anomaly in catchability of the fleets with respect to the fleet /;. The additive catchability model The deterministic model that incorporates the pro- cesses discussed can now be set out. It relies on the changes of the slope [ii i , • ), representing the catchability-at-length pattern, by the addition of the slopes of the partial effects described by Equations 7, 9, and 11. The slope of the catchability model, with the form defined by Equation 4, is represented by p{i.t,EJ) = l3ii.») + p(t) + l3(E) + p{f). 112) The value of /3( ' , /, E, f) represents the slope of the catchability-at-length function without restriction to equilibrium, which has been compensated by the addition of the term fid), where t refers to any ap- propriate unit of time. The red grouper (Epinephelus morio) fishery in the Campeche Bank, Mexico The red grouper on the North continental shelf of Yucatan is exploited by three fleets: an artisanal and a mid-size fleet from Mexico, and a large-scale fleet from Cuba. These fleets overlap in respect to their fishing grounds ( Moreno et al.M: Mexican fleets over- lap in a range of 60*^ to lQ9c\ the midsize fleet from Mexico and large-scale fleet from Cuban, around 609f ; and the artisanal and Cuban fleets, between 12% and 16%. This overlap in fishing gi-ounds means that each fleet causes fishing mortality on different sizes of the stock and thus necessitates that the fleets be consid- ered separately for stock assessment purposes. The artisanal fleet of Mexico is composed of ves- sels of approximately 9 m long, operating from the coast to 20 m depth, and catching mainly juvenile and preadult fish. The midsize fleet uses 15-m ves- sels operating at depths of 10 to 80 m, and catch com- prises both juvenile and adult fish, according to the season. The Cuban fleet fishes mainly with 27-m vessels, operates between 20 and 90 m depth, and catches adult fish. Both Mexican fleets have free ac- cess and move seasonally in the area following changes in fish density (i.e. with reproductive aggre- gation). The Cuban fleet has a catch quota of 3900 t of demersal fish, of which more than 80% is red grou- per. This fleet remains in the central and eastern region of the continental shelf ofYucatan most of the time. Differences related to gear, fishing effort, and fish- ing strategies among fleets are described in Moreno (1980), Burgos (1987), and Fuentes (1987) for the midsize fleet; Saenz et al. ( 1987 ), Salazar ( 1988), and Solana-Sansores and Arreguin-Sanchez ( 1991 ) for the artisanal fleet; Valdes and Padron (1980) for the Cuban fleet; and Seijo (1986), Gonzalez-Cano et al. (1993), and Arreguin-Sanchez et al. (1996) for the three fleets. In this paper we used standardized fish- ing days as the unit of fishing effort. For the catchability model, available data for the red grouper fishery are summarized in Table 2. De- tailed information is derived from the midsize fieet, which harvests around 70% of the total annual yield. Other fleets share the remaining 30% more or less equally. Because the midsize fleet also catches a size range that overlaps the other two, this fleet is the start of our analysis. Later, we will incorporate the other fleets into the catchability model. To solve Equation 3, we assumed that individual growth follows the von Bertalanffy equation with values ofL„ = 87 cm total length (TL) and iC = 0.12/ year (Arreguin-Sanchez-), and the natural mortal- ity coefficient M = 0.3/year (Doi et al., 1981; Contreras etal., 1994). The selection factor for each length class s( ( ), rep- resenting a probability term directly affecting the abundance of length classes in the catch, was set to be s( O = 1. For simplicity, red grouper fishery gears were assumed nonselective (hook-and-line and hand lines with different sizes and baits). Annual catch-per-unit-of-effort( t7)-at-length data for the period 1973 to 1987 for the midsize fleet was used. The iterative procedure described for Equation 3 was applied to each pair of yearly data to obtain initial values of catchability per length class and year. Catchability values for each length class g( /, y) were used to fit the trend with length (Fig. 1) and to de- scribe variability o{ q( i ,•). The solution for Equa- tion 3 does not require a previous knowledge of re- cruits at [/( ' , /) and [/( ' , ^ + 1), and it was solved for Moreno, V., F. Arreguin-Sanchez, M. Contreras, and R. Burgos. 1991. Analysis of usage patterns in shared stocks: the red grou- per fishery from the continental shelf of Yucatan. Mexico. Proc. 44th Ann. Sess. Gulf and Canbb. Fish. Inst., 18 p. IMimeo.l - Arreguin-Sanchez, F. 1996. Length-based growth estimation for the red grouper iEpinephelus morio) in the North continen- tal shelf of Yucatan, Mexico. Centr. Interdiscip. Cienc. Mar., Mexico, 19 p. IManuscript.l 750 Fishery Bulletin 97(4), 1999 Table 2 Summary of the data used in this pape operate with long lines of different sizes Arreguin-Sanchez et al., 1996) r for the red grouper fish (artisanal also uses hook- ery on the north continental shelf of Yucatan, Mexico. Fleets and-line). Fishing effort was standardized as fishing days (see Fleet Annual catch-at-length Monthly catch-at-length Sampling design Site of landing Artisanal' 1987 1987 three stages (port, vessels, Ports along the coast of baskets) Yucatan, Mexico ^-Midsize- 1975 to 1987 (except for 1986) 1975 to 1987 (except for 1986) two stages (day/vessel and Port of Progreso, Mexico commercial size) Cuban fleet- 1975 to 1987 two stages (vessels and Port of La Havana, Cuba commercial size). ' Source: Centro de Investigacion y de Estudios Avanzados del IPN - Source: Institute Nacional de la Pesca. Centro Regional de Invest Yucatan, Mexico. gacion. Pesquera de Yucalpeten. Yucatan. Mexico. n-l or n-2 catchability values, with differences in n caused by the absence of recruitment es- timates. Because we fitted the general tendency as a catcha- bility-at-length pattern, one or two missing points will not af- fect the estimates. Results Length-dependent catchability The fitting process required some consideration of popula- tion behavior because trend could be adjusted to an expo- nential or sigmoidal function. We decided on a sigmoidal form because of reproductive behav- ior. During reproductive aggre- gation, adult fish (>50 cm TL) form groups with a sex ratio (females: males) close to 6:1, comprising individuals of around the same size, but without any apparent size segregation be- tween groups. This is well known by fishermen, and has been described by Moe (1969), Shapiro (1987), and Mexicano-Ci'ntora (1990). In terms of fishing, adults within a specific area have the same prob- ability of catch. In the catchability-at-length trajec- tory, we think this reproductive behavior can be well represented by an asymptotic trend of catchability for large fish. The variation in catchability of young fish, as ex- pressed by one standard deviation (Fig. 1) was simi- 50 60 Total Length (cm) Figure 1 Catchability-at-length pattern for the Mexican midsize fleet of the red grouper fish- ery estimated with Equation :3. The bold line indicates average values over the pe- riod from 197:3 to 1987; thin lines denote one standard deviation, dashed lines rep- resent absolute minimum and maximum values. lar for lengths between 20 to 50 cm and increased sharply after first maturity (at 50 cm TL, as defined by Mexicano-Cintora, 1990). The generalized catchability-at-length equation fitted for the red grouper was £/(/,•): 0.00004256 l-i-e |;j H9(i.(l IH)H-4 ' (1.3) with the standard error (SE) of the ordinate SEa = 0.773, and SE^, = 0.012 for the slope; R- = 0.728 (df=10). Aireguin-Sanchez and Pitcher: Catchability estimates for the Epmephelus mono fishery 751 Time-dependent catchability Time variation of catchability was analyzed considering both annual and monthly catch- structured data. From Equation 7, parameters (Ay) and /3(y) were estimated by regressing Ln[i7(/,n/t/(/,.)] with length (Fig. 2). Table 3 shows values of the constants for both midsize Mexican and Cuban fleets for annual data. On a monthly basis, the same computa- tions were developed for the mid-size fleet. Departures represented in parameters cAm) and l^km ) were estimated in a similar way to those in Equation 7, but considering annual catch-structured data. Seasonal changes in \'ulnerability occurred: q increased in win- ter (December to February) in synchrony with the reproductive aggregation (Fig. 3). Catchability was lower in autumn, when the population is dispersed along the continen- tal shelf and the vulnerability of young fish in coastal waters increases. Catchability differences between fleets Following Equation 11, differences between fleets were evaluated by comparing annual data for the Cuban fleet with those of the mid- size fleet for the period 1975 to 1987. Mexi- can mid-size and artisanal fleets were com- pared on a monthly basis for 1987. Param- eters for a(f) and fiif) were obtained (Tables 4 and 5). Positive values o^ fHf) indicate that relative catchability for small fish is higher for the midsize fleet than for the Cuban fleet; and lower for large fish. This scheme repre- sents differences in fishing areas between fieets. The negative value jiif) for in 1982 suggests fieets changed their usual behav- ior. Similar interpretation can be made for monthly data between the artisanal and mid- size fieets. Figure 3 shows the tendency of Ln[f/ (mid-scale, /,n /[/(Cuban, /,n] ra CO O 2 ' Ot -2 - ■A - 2 - 2 - 1 0 - -1 - 2 0 - -2 - -A - ^juj,,,jjsrrf<^''°^-'\juo^-^~'->'-^^ g^^j;ncxxmnnnnocxx)ocxxoouuuuuuo"c^ 1973 1975 1977 „oO ,0^/-'='='°°°'^'^°°°^^^ -^cneeGOOQQDQooooeeeGo 1979 1981 c^fOr^xjyy:frnMxijJ''^''^'^^ ^p::r^rrx^^C)QiJu^J^J::px>D^^ iHgPOrrooorxJoocxiocujouuuutx/^ oo ^ 1983 1985 1987 20 40 60 80 100 Total Length (cm) Figure 2 Annual tendencies of the ratio Ln[f/(', t}/U{ ',•)! with length for the Mexican midsize fleet of the red grouper fishery representing departure of catchability-at-length pattern with respect to equilib- rium, as expressed in Equation 6. with length. An infiection point is observed just at the length at first maturity (i.e. 50 cm LT). Positive values indicate that q is higher in small (immature) fish for the midsize fieet than for the Cuban fieet, whereas q is equal in both fieets for mature fish. For the tendency of Ln[[/(artisanaU.n/[/( mid-scale, ^n] with length, catchability of immature fish was higher for the artisanal than for the midsize fieet, and again equal for adult fish up to 65 cm TL. For fish larger than 65 cm, q was higher in the midsize fieet. 752 Fishery Bulletin 97(4), 1999 Ln U(mid- size, ^,y) U(Cuban,^,y) 40 50 60 Total Length (cm) 80 90 Figure 3 Trend in catchability-at-length differences, between Mexican midsize and Cuban fleets for the period 197.5 to 1979, as expressed by Equation 11. Positive values indicate higher vulnerability for immature fish for the midsize fleet than for the Cuban fleet. Vulnerability of adult fish to both fleets is equal. Table 3 Parameters of Equation 7 for the Mexican midsize of recruits each year and /3ly) departure of catcha and Cuban fleets for the red grouper fishery. a(y) bility trend with length, both with respect to eq expresses relative vu uilibnum. (*P<0.05; nerability **P<0.01). Year a(y) midsize midsize i?2 My) Cuban Ry) Cuban R^ 1973 -0.509 0.010 0.405* 1974 0.508 -0.013 0.664* 1975 0.511 -0.010 0.487** -1.494 -0.018 0.602* 1976 0.953 -0.025 0.610** 0.865 -0.006 0.507** 1977 0.683 -0.022 0.600** 0.767 0.024 0.624* 1978 1.045 -0.024 0.686** 0.416 0.047 0.496* 1979 0.941 -0.021 0.757* -1.396 0.033 0.771** 1980 -0.612 0.015 0.614** -2.925 -0.043 0.544** 1981 0.476 -0.013 0.643** -2.147 -0.026 0.622* 1982 0.487 -0.012 0.396* 1.923 -0.021 0.483** 1983 -0.401 0.009 0.371* 1.448 0.020 0.701** 1984 0.175 -0.005 0.446** 0.828 0.079 0.513** 1985 0.135 -0.004 0.432* -1.126 0.026 0.661** 1986 -1.064 0.023 0.739* -4.889 0.079 0.596** 1987 -0.915 0.013 0.581* -1.2,50 0.026 0.470** Amount of fishing and density-dependent catchability As in Equation 9. values of fHy) from Table 3 were fitted against fishing effort at time t. Figure 4 shows the tendency for both the midsize and Cuban fleets. For the midsize fleet, the model gives the following values for the parameters: aiE) = 0.1252, (iiE) = 4.8 • lO-*^; R~ = 0.74. Although the tendency for the Cuban fleet was also negative, the observed variability was Arreguin-Sanchez and Pitcher: Catchability estimates for the Epinephelus mono fishery 753 higher, and at least two prob- able tendencies were empiri- cally identified (Fig. 4B), both of them as a consequence of changes in the catch quota as- signed to the Cuban fleet. For further computations, the den- sity-dependent effect was as- sumed to be similar for both fleets because of the absence of more detailed information to support quantitative analysis. This assumption is reasonable because both fleets average more than 60*^ overlap in a year, but almost 100'7( during reproductive aggregation. Because of the evident differ- ences in behavior between juve- nile and reproductive fish, the density-dependent effect of the amount of fishing was also ap- plied within each length class, as in Equation 9. Parameters of the model are given in Table 6. Variation of the slope ji(E) with length class is shown in Figure 5. Once again, immature fish behaved differently from adults. The density-dependent effects for immature fish were con- stant, whereas the same effect increased with size for sizes >50 cm TL. It is clear that the den- sity-dependent effect during ag- gregation promotes a greater catcha- bility, and because the older adult fish tend to remain longer in deeper waters (Shapiro, 1987), the density- dependent effect will increase with size (age). The additive catchability model for the red grouper fishery Once several sources of variation were estimated, a model (Eq. 12) was obtained by adding partial effects to the slope of Equation 13, as follows: m 20 22 24 26 28 mid-size fleet 30 32 m 0.09 0.04 -0.01 -0.06 B 86 "•«^- /■■■as ••-78 i85 I /-y- - . 76 81 • 82 '• ♦■75. Cuban fleet 13 15 " ■ - .• 80 9 11 Fishing effort (10' days) Figure 4 Estimation of density-dependent effects on catchability-at-length pattern for (A) the Mexican midsize fleet and (B) Cuban fleet of the red grouper fishery (Eq. 111. General trends suggest density-dependent effects by fleets are of the same magni- tude (see text for interpretation!. q(i,t,f,E}: where 0.00004256 -{3.896* /J(',ly.ml./\£)i} 1+e P(E) 2 - 20 30 40 50 60 70 80 90 Total Length (cm) Figure 5 Ti'end in density-dependent effects on q within each length class with size. 754 Fishery Bulletin 97(4), 1999 Table 4 Estimates of parameters for Equation 11 describing de- parture of catchability coefficient with length for the Cu- ban fleet with respect to the Mexican midsize fleet for the red grouper fishery (*P<0.05; **P<0.01). Year a(f) P(f) R^ n 1975 -4.474 0.052 0.636** 14 1976 -2.956 0.047 0.613* 13 1977 -3.590 0.058 0.807** 14 1978 -3.904 0.065 0.822* 12 1979 -5.186 0.076 0.795* 11 1980 -4.713 0.055 0.736* 12 1981 -6.476 0.097 0.708* 13 1982 0.323 -0.020 0.711** 9 1983 -4.527 0.051 0.713* 10 1984 -3.936 0.059 0.561** 11 1985 -3.847 0.050 0.848** 13 1986 -6.149 0.070 0.873* 13 1987 -3.942 0.055 0.606* 11 l3((,y,mJ\E) = li(i,») + [p{y),p{iu)] + li{f) + (HE), where / indexes fleet, and each constant takes the corresponding value, depending on the specific length class ( ' ), time, year (_v) or month (m I, fleet if), and the specific density-dependent effect {E}. Discussion The catchability model incorporates several sources of variation, such as individual size, time variations, and density-dependent effects, for which we assumed fishing effort is the most important variable affect- ing fish density, and the participation of several fish- ing fieets which differ in gear, catching power, and effort allocation. The additive-catchability model, as applied to the red grouper, performed well and allows us to iden- tify some aspects of fish behavior that influence catchability. For example, reproductive aggregation increases vulnerability offish. Of particular impor- tance are the relative constancy of catchability and its variance for immature individuals, and the in- creasing vulnerability of adult fish with size, espe- cially during reproductive aggregation. This is re- lated to the reproductive behavior of the population. Young fish remain close to the coast, with east-west seasonal displacements, whereas adults move across the continental shelf and aggregate during winter for reproduction. This means that juveniles and Table 5 Estimates of parameters for Equation 11 describing de- parture of catchability coefficient with length for the mid- size fieet with respect to the artisanal fleet, both from Mexico (*P<0.05; **P<0.01). Month a(.f) Pin R^ n January 2.953 -0.071 0.576* 23 February 2.654 -0.083 0.701** 22 March 4.180 -0.107 0.719* 24 April 7.904 -0.230 0.965** 14 May 2.524 -0.064 0.769* 28 June 1.295 -0.032 0.483* 31 July 1.426 -0.041 0.720* 28 August -1.589 0.041 0.562** 16 Table 6 Estimates of parameters for Equation 9 represent ing den- sity-de pendent effects with in each length class (h = 13, | dr=ll; ■P<0.05; **P<0.01). Total Standard Standard length aiE) error piE) error (cm) io-« a(£)10-" io-» /3(E)10-9 R' 25 3.76 1.211 1.238 1.282 0.89** 30 31.05 3.356 1.164 3.471 0.50'* 35 54.72 3.276 1.013 2.905 0.52** 40 48.32 1.574 0.926 2.013 0.66* 45 26.39 1.147 1.237 2.718 0.65' 50 3.68 1.091 1.936 2.426 0.85* 55 -0.31 1.123 2.360 2.974 0.85* 60 0.62 1.144 2.678 2.857 0.89* 65 1.56 1.316 3.198 1.892 0.96** 70 -1.17 1.388 3.886 2.593 0.95** 75 -2.83 1.510 4.589 2.621 0.96** 80 0.43 0.743 5,549 3.282 0.96* adults respond differently to fieets. The model also identified the differential accessibility of specific size ranges to different fieets. This is important because, in a fully exploited fish resource such as the red grou- per, controls on fishing mortality must enhance sur- vival of spawners and recruits. Figure 6 summarizes the relation between fishing effort, maturity, and seasonal changes in catchability for the red grouper fishery. During reproductive aggregation, identified by the maturity index, vulnerability increases and vessels spend less fishing effort to obtain a given level of yields. After reproduction, groupers disperse along the continental shelf and fleets must increase fish- Arreguin-Sanchez and Pitcher Catchability estimates for the Epinephelus mono fishery 755 ing intensity because vulnerability offish will decrease. For the density-dependent catchability, the model describes fish behavior reasonably well. The red grouper is a gregarious and territorial fish. Stock density is reflected straightfor- wardly in yields. If fish densities decrease, q will decrease, and vice versa. This response to the amount of fishing is reflected by the model. For the catchability model, note that al- though a transition matrix is very helpful for the estimation process, the form of the catchability-at-length function must be inter- preted with good understanding of the biology of the exploited fish resource and of the fish- ery. Fish biology probably is the most important aspect because the effect of other sources of varia- tion are added to the slope of this function. The resulting catchability model can be eas- ily incorporated into estimations of fishing mortality and population size; i.e. for the mid- size fleet at time /, fishing mortality can be expressed by Jan -0 05 0 00 0,05 Catcfiability deviation 0.05 015 Fishing effort index t- - ■ -A 0 0 0 1 0 2 H/latunty index Figure 6 Schematic representation of the main q features of the red grou- per fishery on the Campeche Bank: l Al seasonal catchability varia- tion represented as departure of the annual average; (B) fishing effort index, represented as a monthly proportion of the year; (Cl maturity index, reflecting diameter of oocytes. F(^^midsize,£') (/( ^^midsize,£) £'(^, midsize) (14) For the population size, the participation of the three fleets can be incorporated. A^ ■I ; = 1 f- 1 F ^ (=1 y C 1 3 "l V 1 = 1 ) ( '^ ] -Lm*f 'j.ir,.E]\ l-e • ' '-' '^ \ 1=1 / (15) Despite the potential use of the catchability coeffi- cient in some stock assessment models (i.e. age-based virtual population analysis, VPA, Pope, 1972; and length-based VPA, Jones 1981, 1984; Gulland^), Equations 14 and 15 suggest an alternative stock assessment tool, based on length-composition data. A detailed description of catchability through the additive model reflecting fish behavior and fishei-y practices could be very useful for management pur- poses when it is incorporated into assessment mod- Gulland, J. A. 196.5. Estimation of mortality rates. Annex to rep. Arctic Fish. Working Group. ICES Council Meeting (CM I, 196.5. paper .3, 9 p. els such as the above. Currently the red grouper fish- ery is subjected to heavy fishing where the main prob- lems are intensive fishing of juveniles and the high vulnerability of adults. Because the additive catcha- bility model describes these processes and incorpo- rates all fleets, results can strongly aid management decisions based on a differential control of fishing mor- tality within the stock structure and between fleets. Even when the model can describe the study well, we observed some critical aspects during application. The first is that we must know red grouper behavior sufficiently well to decide the form of the catchability- at-length relation (Eq. 4). Statistical criteria (i.e. a correlation coefficient) are not sufficient. A wrong in- terpretation of this could mean a disaster for the fish- ery. A second and similar aspect can occur for den- sity-dependent catchability. The third aspect is the significance of sources of variation. Because the catchability model is based on the addition of the slopes of each particular variable and relation (i.e. with length, time, etc. ), we must test that these slopes are significantly different from zero (see Equations 4, 7, 9, and 11), and if they are not different, the source of variation tested does not help to explain changes in q. The fourth aspect involves the number of para- meters to be estimated, which usually is taken as proportional to the complexity of the model compu- tations, and inversely related to applicability. For the present case, the number of parameters in the model is /; + 2, where n is the number of sources of varia- 756 Fishery Bulletin 97(4), 1999 tion to be considered. However, the number of con- stants to be estimated strongly increases with the length of the time-series of data as 3im»yf) + 2( i ) + 2, with m = number of months, v = number of years, /"^number of fleets, and / = number of length classes. For example, in our study case, with v = 13,/= 2, and ( = 12, the number of constants to be estimated will be 104. We think these are a lot of parameters; however, they are easily estimated because all of them are ob- tained by regression or by an analogous technique. Acknowledgments Some of the initial ideas developed in this paper were inspired by the late John A. Gulland, while F. A.-S. was at the Renewable Resources Assessment Group, London, in 1989. We want to dedicate this contribu- tion to his memory. Our gi-atitude is also extended to S. Rodriguez, L. Capurro, and D. Pauly for their sup- port and valuable comments on an early draft. We also thank three anonymous referees for their valu- able comments. The first author was funded by the National Polytechnic Institute through COFAA, EDD and DEPI. Lastly, we thank Ellis Glazier, CIBNOR, who edited the English-language text. Literature cited Arreguin-Sanchez, F. 1996. Catchability: a key parameter for fish .stock assess- ment. Rev. Fish Biol. Fish. 6:1-22 Arreguin-Sanchez, F., M. Contreras, V. Moreno, R. Burgos, and R. Valdes. 1996. Popuhition dynamics and stock assessment of the red grouper tEpinephelus morio) fishery on Campeche Bank, Mexico. In F. Arreguin-Sanchez, J. L. Munro, and D. Pauly (eds.i. Biology of tropical groupers and snappers. ICLARM Conf Proc. 48:202-217. Burgos, R. 1987. Operaciones y rendimientos de la flota pesquera mayor de Yucatan durante 1985. Ph.D. diss., Contr. In- vest. Pesq. Cent. Reg. Invest. Pesq. Yucalpeten, Inst. Nal. Pesca, Mexico. Caswell, H. 1988. Approaching size and age in matrix population models. In E. Ebenman and L. Persson leds. I, Size-struc- tured populations, 8.5-105 p. Springer- Verlag, Berlin, Heidelberg. Contreras, M., F. Arreguin-Sanchez, J. A. Sanchez, V. Moreno and M.A. Cabrera. 1994. Mortality and population size of the red grouper iEpinephelus morio) fishery from the Campeche Bank, Mexico. Proc. 4.3th Ann. Scss. Gulf Caribb. Fish. Inst., .39.3-401 p. Csirke, J. 1988. Small shoahng pelagic fish stocks. In .]. A. Gulland (ed.). Fish population dynamics, 2nd ed., p. 271-.302. John Wiley and Sons, London. 1989. Changes in the catchability coefficient in the Peru- vian anchoveta tEngraulis ringens) fishery. In D. Pauly, P. Muck, J. Mendo, and I. Tsukayama (eds.). The Peruvian upwelling ecosystem: dynamics and interactions, 207-219 p. ICLARM Conf Proc. 18. IMARPE-Peru, GTZ-Germany, ICLARM-Philippines. Doi, T., D. Mendizabal, and M. Contreras. 1981. Analisis preliminar de la poblacion de mero Epinepheliis mono (Valenciennes) en el Banco de Cam- peche. Ciencia Pesquera, Inst. Nal. Pesca., Mexico. 1(1): 1-16. Fuentes, D. 1987. La pesqueria de mero en el Banco de Campeche. Mem. VII Congr. Nal. Oceanogr. Ensenada, Mexico 1:361- 374. Gonzalez-Cano, J., F. Arreguin-Sanchez, M. Conteras, V. Moreno, R. Burgos, C. Zetina, and V. Rios. 1993. Diagnbstico del estado de la pesqueria de mero {Epinepheliis mono) en el Banco de Campeche. Informe del Grupo de Trabajo sobre el Recurso Mero. Inst. Nal. Pesca. Mexico, 47 p. Grecco, V., and W. J. Overholtz. 1990. Causes of density-dependent catchability for George's Bank haddock Melanogrammus aegelfinus. Can. J. Fish. Aquat. Sci. 47:.385-394. Gulland, J. A. 1983. Fish stock assessment: a manual for basic methods. .John Wiley and Sons, New York, NY, 223 p. Hilborn, R., and C. J . Walters. 1992. Quantitative fisheries stock assessment: choice, dy- namics and uncertainty. Chapman and Hall. New York, NY, 570 p. Jones, R. 1981. The use of length composition data in fish stock as- sessment (with notes on VPA and cohort analysis). FAO Fish. Circ. 734, 55 p. 1984. Assessing the effect of changes in exploited pattern using length composition data. FAO Fish. Tech. Pap. 256, 118 p. MacCall.A. D. 1976. Density dependence of catchability coefficient in the California Pacific sardine, Sardinops sagax caerulea, purse seine fishery. Calif Coop. Oceanic Fish. Invest. Rep. 18:136-148. Mexicano-Cintora, G. 1990. Analisis preliminar de algunos aspectos reproductivos del mero iEpinephelus morio) de las costas de Yucatan. Rep. Esp. Acad. Centr. Invest. Est. Avanz. Inst. Politec. Nal., Merida, Mexico. 42 p. Moe, M. A. 1969. Biology of the red grouper Epinephel iis mono (Valenciennes) from the eastern Gulf of Mexico. Fla. Dep. Nat. Res. Prof Ser. 10, 95 p. Moreno, V. 1980. La pesqueria de mero iEpinephelus mono) en el Estado de Yucatan. Ph.D. diss. LIniv. Auton. Edo. Morelos. Mexico, 63 p. Murphy. ('•■ 1966. Population biology of the Pacific sardine iSardinops caerulea I Proc. Calif Acad. Sci. 34:1-84 Pitcher, T. J. 1996. The impact of fish behavior on pelagic fish and fisheries. Scicntia Marina 59(3-4):295-306. Pope, J. G. 1972. An investigation of accuracy of Virtual Population Analysis. Int. Comm. NW Atl. Fish. Res. Bull. 9:65-74 ArreguinSanchez and Pitcher: Catchability estimates for the Epinephelus mono fishery 757 Quinn, T. J. 1987. Standardization of catch per unit effort for short-term trends in catchability. Nat. Res. Model. 1:279-296. Rose, G. A., and W. C. Legget. 1991. Effects of biomass-range interactions on catchability of migratory demersal fish by mobile fisheries; an example of Atlantic cod [Gadus morhua). Can. J. Fish, aquat. Sci. 48: 843-848. Saenz, M., F. Mendoza, and J. C. Piste. 1987. Diagnosis de la pesqueria artesanal del Estado de Yucatan. Contr. Invest. Pesq. Cent. Reg. Invest. Pesq. Yucalpeten. Inst. Nal. Pesca. Mexico. Dcto. Tec. No. 5:1-18. Salazar, A. R. 1988. Contribucion al conocimiento de la pesqueria de mere ^Epinephelus mono) de la flota menor de las costas de Yucatan. Tesis Lie. Esc. Nal. Ens. Prof Iztacala, Univ. Nal. Auton. Mexico, Mexico. 81 p. Seijo, J. C. 1986. Comprehensive simulation model of a tropical dem- ersal fishery: red grouper (Epinephelus morio) of the Yucatan continental shelf Ph.D. diss., Michigan State Univ. city, MI, 210 p. Shapiro, D. Y. 1987. Reproduction in groupers. In J. J. Polovina and S. Ralston, (eds.). Tropical snappers and groupers: biology and fisheries management, 295-328 p. Westview Press, London. Shepherd, J. G. 1987. Towards a method for short-term forecasting of catch rates based on length compositions. In D. Pauly and G. R. Morgan, (eds.), Length-based methods in fisheries re- search. ICLARM. Manila; and KISR, Kuwait. ICLARM Conf. Proc. 14:167-176. Solana-Sansores, R,, and F. Arreguin-Sanchez. 1991. Diseho de muestreo probabilistico para la pesqueria artesanal de mero (Epinephelus morio) del Estado de Yucatan, Mexico. Ciencias Marinas 17(l):51-72. Valdes. E., and G. Padron. 1980. Pesquerias de palangre. Revista Cubana de Investi- gaciones Pesqueras. 5(2);38-52. 758 AbStrSCt.— After spending summer months in estuaries, spring- and sum- mer-spawned young-of-the-year (YOY) bluefish, Pomatomus saltatrix. migrate out to continental shelf waters of the Mid-Atlantic Bight in early autumn. Adult bluefish are found on the conti- nental shelf throughout summer and fall. Both juveniles and adults have high food consumption rates and are gener- ally piscivorous. To determine princi- pal prey types on the shelf, dietary analyses were performed on YOY and adult bluefish collected from National Marine Fisheries Service autumn bot- tom trawl surveys in 1994 and 199.5. Both spring- and summer-spawned YOY bluefish diets were dominated by bay anchovy. However, the significantly larger size of the spring-spawned cohort was associated with the consumption of other prey species such as squid, butterfish. striped anchovy, and round herring. Summer-spawned bluefish were significantly smaller in 199.5 than in 1994; diet and prey size comparisons suggest that body size had a dramatic influence on the amount of piscivorous feeding in the summer-spawned cohort. Adult bluefish diet was dominated by schooling species such as squid, butter- fish, and clupeids. Cannibalism was virtually nonexistent. Daily ration es- timates of YOY bluefish on the shelf (4- 12"^? body wt/d i were similar to estua- rine estimates in late summer It is es- timated that during the month of Sep- tember, YOY bluefish in aggregate con- sumed 6.0 to 6.8 billion bay anchovies in 1994 and from 2.2 to 5.3 billion in 1995. The effect of this predatory loss on population dynamics of bay anchovy and the fish community on the conti- nental shelf is unknown. Foraging habits of bluefish, Pomatomus saltatrix, on the U.S. east coast continental shelf* Jeffrey A. Buckel Marine Sciences Research Center State University of New York Stony Brook, New York 11794-5000 Present address James J Howard Manne Sciences Laboratory National Marine Fishenes Service, NOAA 74 Magruder Road Highlands, New Jersey 07732 Email address ibuckelgsh nmfsgov Michael J. Fogarty Northeast Fisheries Science Center National Manne Fisheries Service, NOAA 166 Water Street Woods Hole, Massachusetts 02543 David O. Conover Marine Sciences Research Center State University of New York Stony Brook, New York 11794-5000 Manuscript accepted 1 December 1998. Fish. Bull. 97:758-775(1999). One of the dominant marine pisci- vores along the U.S. east coast is the bluefish, Pomatomus saltatrix (Juanes et al., 1996). Bluefish spawned in offshore waters of the South Atlantic Bight in the spring (spring-spawned ) are advected north- ward in waters associated with the Gulf Stream (Hare and Cowen, 1996) and move into mid-Atlantic Bight estuaries in June at -60 mm fork length (Kendall and Walford, 1979; Nyman and Conover, 1988; McBride and Conover, 1991). A second wave of recruits consisting of summer- spawned fish occur in nearshore wa- ters from mid to late summer. In estuaries, bluefish exhibit rapid growth rates that are fueled by high food consumption and evacuation rates (Juanes and Con- over, 1994; Buckel et al., 1995; Buckel and Conover, 1996). In the Hudson River estuary, for example, preda- tion by bluefish is high enough to account for virtually all natural mor- tality of young-of-the-year (YOY) striped bass during their summer- fall growing season (Buckel et al., 1999). After spending the summer months in estuaries, YOY bluefish migrate back out onto the shelf and migrate southward to overwinter (Munch and Conover, in press). Spring-spawned YOY bluefish mi- grate out of estuaries at a size of -180 mm fork length (FL) and -100 g ( Nyman and Conover, 1988; McBride and Conover 1991). By age 1, these spring-spawned fish attain sizes of -260 mm FL and weights of -300 g (Chiarella and Conover, 1990). Therefore, an individual spring- spawned bluefish gains from 100 to 200 g during its autumn migration. Given a 15% gross growth efficiency (Juanes and Conover, 1994; Buckel et al. 1995), a single bluefish could thus consume from -650 to 1300 g of prey ■ Contribution 1134 of the Marine Sciences Research Center, State University of New York, Stony Brook, New York 11794. Buckel et al.: Foraging habits of Pomatomus sa/tatrix 759 Compared with the nearshore phase of the early life history of bluefish, our knowledge of the forag- ing ecology and predatory impact of bluefish on the continental shelf is poor. Here we quantify the diet of YOY and adult bluefish and determine YOY blue- fish prey type and size selectivity patterns, foraging chronology, daily ration, and biomass of prey con- sumed during the autumn migration on the shelf. Methods Study area and collections YOY and adult bluefish and their potential prey were collected on the U.S. east coast continental shelf dur- ing the autumn of 1994 and 1995 aboard the research \essel Albatross IV. Collections were made during the National Marine Fisheries Service, Northeast Fisheries Science Center's (NMFS-NEFSC) autumn bottom trawl survey cruises at predetermined sta- tions from Cape Hatteras, NC, to Nova Scotia. Cruises began in early September and ended in late October in both years (6 September to 28 October 1994 and 5 September to 27 October 1995). Descriptions of the survey design and the trawl characteristics can be found in Azarovitz (1981). Briefiy, tows were made with a no. 36 Yankee trawl equipped with rollers with an opening 3.2 m high, 10.4 m wide, with 12.7-cm stretched mesh in the opening, with 11.4-cm stretched mesh in the codend, and with a 1.25-cm stretched mesh lining in the codend and upper belly to retain YOY fishes. Tows were 30 minutes in duration at 3.5 knots in relation to bottom and were conducted on a 24-h basis. Dietary analyses Adult and YOY bluefish were distinguished by us- ing length-age relationships. Munch and Conover ( in press) examined the annual size distributions of blue- fish from autumn bottom trawl surveys conducted since the 1970s in four different regions of the shelf: SOC=South of Chesapeake Bay; C-D=Chesapeake Bay to Delaware Bay; SNE=Southern New England (Delaware Bay to Narragansett Bay); and Georges Bank. On the basis of these size distributions and backcalculated sizes at age 1 (Chiarella and Conover, 1990, and references therein), they classified YOY bluefish in autumn as those fish <300 mm FL. We classified adults as those bluefish >300 mm FL and we distinguished spring- and summer-spawned YOY bluefish for each region in 1994 and 1995 on the ba- sis of bimodality in length-frequency distributions (Munch and Conover, in press). We also adopted Munch and Conover 's (in press) geographical bound- aries for this analysis. In 1994, summer-spawned blue- fish sizes by region were as follows: SOC, FL<120 mm; C-D, FL<160 mm; and SNE, FL<160 mm. In 1995, summer-spawned bluefish sizes were as follows: SOC, FL<100 mm; C-D, FL<150 mm; and SNE, FL<140 mm. Spring-spawned bluefish were those fish larger than the summer-spawned cohort by re- gion but <300 mm FL. Diets of spring- and summer-spawned juveniles and adult bluefish were quantified. Bluefish taken for stomach content analysis were wet weighed (±1.0 g) and measured for fork length, FL (to 1.0 mm). Stom- achs were removed at sea and preserved in 10% for- malin buffered with sea water. On some occasions, whole fish were either frozen or presei-ved in lO'^ for- malin and then processed in the laboratory. Stomach contents of bluefish were identified to the lowest pos- sible taxon, enumerated, blotted dry, weighed (±0.01 g), and measured ( TL for fish prey and mantle length for squid, ±1.0 mm). Eye diameter (±0.1 mm) and caudal peduncle height (±0. 1 mm ) were measured for partially eaten bay anchovy and butterfish prey and converted into TL from linear regression equations (Scharfet al., 1997;Scharfetal., 1998a). Each trawl containing bluefish provided us with a group or "cluster" of bluefish for a given station. Therefore, for our analysis, the mean and variance of diet indices were calculated with cluster sampling estimators developed at the NEFSC-NMFS, Woods Hole, MA (Cochran, 1977, Fogarty, unpubl. data). These calculations are described in a previous study that examined bluefish diet in the Hudson River es- tuary (Buckel et al., 1999). Net feeding During a 30-minute tow, bluefish may feed within the trawl (an activity known as net feeding). Net feed- ing could bias the diet index estimates or affect gut fullness level estimates (or do both). We tested for such bias by classifying fish or squid prey as either 1) "fresh" or 2) "digested" during our stomach con- tent examination. "Fresh" prey had no sign of diges- tion and "digested" prey were either partially or well digested (e.g. they were anywhere from starting to lose skin to being only identifiable by skeleton or shape). The percent occurrence of four dominant prey (bay anchovy, striped anchovy, butterfish, and squid) were compared between fresh and digested catego- ries in YOY bluefish. If the percent occurrence of each prey category within a cohort and region was simi- lar for fresh and digested prey, it would suggest that either there was no net feeding or that it did not af- fect diet index estimates. 760 Fishery Bulletin 97(4), 1999 Prey-type selectivity The feeding selectivity of spring-spawned bluefish on the continental shelf was determined from the relative abundance of prey in individual bluefish guts and the relative abundance of prey in trawl catches. Four prey categories were examined in 1994 (bay anchovy, butterfish, squid, and "other") and three prey categories in 1995 (bay anchovy, squid, and "other"). The "other" category included any teleost fish that did not fall into the previously mentioned groups. The relative abundance of these prey in the field was calculated by including only those prey that were of a size that bluefish could theoretically con- sume (prey FL<80'^ of bluefish PL; determined from an independent study, Scharf et al., 1997). Addition- ally, the index was calculated for stations where at least ten spring-spawned bluefish were captured. Chesson's (1978) index was used to determine blue- fish prey preference as a. r, /P, i=L ,m, where a, P, = m the selectivity for prey type / for an in- dividual bluefish; the relative abundance of prey type / in an individual bluefish stomach; the relative abundance of prey type / in the environment; and the number of prey types available. Values of a, were averaged for each year. Random feeding occurs when mean a= llm (1994=0.25 and 1995=0.33); values of a^ > llm or a^ < llm represent "selection" and "avoidance" of prey, respectively. Ran- dom feeding was tested by using a ^-test to compare mean a, with 1//?; for each prey type within each year (Chesson, 1983). Prey-size relationships and size selectivities Prey sizes of all bluefish prey wei'e measured directly or determined indirectly from regression equations. The relationship between ingested prey size and blue- fish length was determined for YOY and adult blue- fish in both 1994 and 1995. In order to compare the relative prey size ingested by spring- and summer- spawned bluefish cohorts in 1994 and 1995, the fre- quency distributions of bay arichovy FL to bluefish FL ratios were examined. Ratios were calculated for all bay anchovy prey that were measured ( 1994, n=388; 1995, ;i=202). Size-selective feeding on bay anchovy prey was examined by comparing the sizes of bay anchovy in- gested by bluefish at a specific station with sizes of bay anchovy captured in the trawl at that station. Stations from 1994 and 1995 that had «>15 bay an- chovy length measurements for each bluefish cohort were used in the analysis. Bay anchovy lengths at each station were obtained from archived data at NEFSC-NMFS, Woods Hole, MA. Selectivity for bay anchovy prey size categories was measured by using Chesson's ( 1978 ) index ( described above). Bay anchovy lengths ingested by bluefish and collected at each station were partitioned into four bay anchovy FL categories: 25-34, 35-44, 45-54, and 55-64 mm. Values of a, were calculated from indi- vidual bluefish for each size bin at each station sepa- rately; these were then averaged across stations. As above, values of a, > 1/m or a^K llm represent "selec- tion" and "avoidance" of size categories, respectively. Estimated values of mean a, were compared with llm by using a Mest ( 1994 spring-spawned, lA?i=0.25, and summer-spawned, l/«z=0.33; 1995 spring- spawned, l/m=0.33). Feeding chronology, daily ration estimates, and impacts on bay anchovy Values of gut fullness were used to examine feeding chronology of YOY bluefish. Gut fullness values (F) were calculated as F=GIW, where G = prey wet weight; and W = bluefish wet weight (total weight minus prey wet weight). Gut fullness values for individual bluefish were pooled over eight 3-h time periods and averaged. This analysis was performed for both spring- and sum- mer-spawned bluefish by geographic region. In order to estimate feeding rates of bluefish on the shelf, bluefish gastric evacuation rate (GER) es- timates were needed. Previous work on YOY blue- fish GER showed that prey type and bluefish body size did not have a significant effect on bluefish GER; of the factors examined, temperature had the only sig- nificant effect on bluefish GER (Buckel and Conover, 1996). These laboratory experiments described blue- fi.sh GER from 21 to 30C. In our continental shelf collections, YOY bluefish were found in water tempera- tures as low as 15^C (mean of surface and bottom tem- perature). Therefore, a laboratory experiment to mea- sure YOY bluefish GER at 15''C was performed. The experiment was conducted in an identical manner to previous GER experiments on YOY blue- Buckel el al.: Foraging habits of Pomatomus saltatnx 761 fish (Buckel and Conover, 1996). Brieny, YOY blue- fish were acclimated to experimental tanks for two weeks, acclimated to 15°C for 3 days, and then held at IS'C for a 12-48 h starvation period. They were then fed a single meal of previously frozen and thawed adult bay anchovy (bluefish sizes were: mean bluefish TL=122 |range=99-146|, mean wet weight=15.79 g, prey mean wet weight=0.88 g, mean prey wt/predator wt=5.8%). After a predetermined period of time, indi- \idual bluefish were sacrificed and their stomach con- tents removed, blotted dry, and weighed (±0.01 g). The exponential GER model (see Buckel and Conover, 1996) was fitted to the proportion of meal remaining versus time by using nonlinear regi'ession analysis. The GER estimate from this experiment was used along with estimates from Buckel and Conover ( 1996) to develop an empirical function de- scribing YOY bluefish GER from 15° to 30"C. This function was used to estimate bluefish GER from water temperatures at which bluefish were collected on the continental shelf. Bluefish daily ration was estimated for each cohort by region in 1994 and 1995 with the Eggers ( 1979) equation: D = F-R2A, YOY bluefish abundance in 1994 and 1995 were used to calculate the daily consumption of bay anchovy by the YOY bluefish population in the following steps. First, the biomass of the spring- and summer- spawned cohorts on the shelf was calculated. This was done by determining the numbers in each spawn- ing cohort (this was necessary because the VPA was for all YOY combined) based on relative abundance estimated from the NMFS autumn bottom trawl sur- vey and then multiplying by the average individual fish weight in each cohort. Secondly, this cohort bio- mass was multiplied by the estimates of daily ration (lowest and highest) to determine the daily biomass of prey consumed by each cohort. To determine the amount of bay anchovy consumed daily, each cohort's biomass was multiplied by the mean proportion of bay anchovy by weight in each cohort's diet. Finally, a daily estimate of the numbers of bay anchovy con- sumed by the YOY bluefish population was calcu- lated by dividing the biomass of bay anchovy con- sumed by the average bay anchovy weight ingested by each cohort. Bay anchovy weight was calculated from mean bay anchovy length ingested with regres- sions from Hartman and Brandt ( 1995a >. where D = is bluefish daily ration; F, = the mean gut fullness over 24 hour; and R^, = the exponential gastric evacuation rate (see above). Mean gut fullness was calculated by taking the av- erage of the individual time period (/) gut fullness means from the feeding chronology analysis (see above). The mean of the means was used to give each time period equal weight even though sample sizes for given time intervals varied over the diel cycle. Daily ration was calculated for those geographical regions and cohorts that had a sufficient diel record (e.g. gut fullness estimates throughout the diel cycle). The standard eiTor of the daily ration estimate was approximated by using the delta method ( Seber, 1973 ). Estimates of the biomass of prey consumed by the YOY bluefish population during their southward migration requires estimates of the numbers of YOY bluefish in the population. The NEFSC, NMFS, Woods Hole, MA, has used virtual population analy- sis (VPA) to estimate the abundance of the east coast bluefish population (NEFSC). These estimates of NEFSC. 1997. Report of the 23rd Northeast Regional Stock Assessment Workshop (23rd SAW): Stock Assessment Review Committee (SARC) consensus summary of assessments. North- east Fisheries Sci Cent. Ref. Doc. 97-05, 191 p. I Available from National Marine Fisheries Service, 166 Water Street, Woods Hole, MA 02543-1026.1 Results Dietary analyses The stomach contents of 989 young-of-the-year (626 spring- and 363 summer-spawned) and 275 adult bluefish were examined. All YOY bluefish were cap- tured between 12 and 29 September 1994 and 7 and 23 September 1995. Continental shelf bluefish diets were dominated by teleost fish and squid prey (see Table 1 for common and scientific names of bluefish prey). In both 1994 and 1995, the dominant fish prey of spring-spawned bluefish in all three geographical regions was bay anchovy (Table 2). Other spring- spawned bluefish prey included long-finned squid, striped anchovy, butterfish, and round herring. But- terfish were slightly more important in 1994 than in 1995. Also in 1994, amphipods made up a substan- tial portion of spring-spawned bluefish diets in the SNE region. In 1995, channeled whelk were a rela- tively important prey of bluefish in the C-D region; the foot and operculum were found in bluefish col- lected over a large geographical area. The dominant fish prey of summer-spawned blue- fish across all geographical regions was bay anchovy in both 1994 and 1995 (Table 3). The incidence of long-finned squid in summer-spawned bluefish di- ets was low; however, other invertebrates such as 762 Fishery Bulletin 97(4), 1999 amphipods, mysids, crab larvae, and copepods were relatively important prey. This was particularly true for the relatively small size summer-spawned blue- fish collected in the SNE in 1995; the diet of these fish were dominated by copepods (Table 3). The diet of adult bluefish collected in the Georges Bank region was dominated by butterfish, squid, round herring, and Atlantic herring in both 1994 and Table 1 Common and scientific names of bluefish prey items on the continental shelf and Georges Bank. Common name Scientific name Fish American plaice Hippoglossoides plattessoides Atlantic cod Gadus morhua Atlantic herring Clupea harengus Atlantic mackerel Scorn ber scorn brus Atlantic silverside Menidia menidia bay anchovy Anchoa mitchilli butterfish Peprilus triacanthus conger eel Conger oceanicus fawn cusk eel Lepophidium cervinum fourbeard rockling Enchelyopus cmjbrius haddock Melanogrammus aeglefinus Northern pipefish Syngnathus fuscus Northern puffer Sphoeroides maculatus ocean pout Macrozoarces americanus planehead filefish Monacanthus hispidus red hake Urophycis chuss round herring Etrumeus teres sand lance Ammodytes spp. soup Stenotomus chrysops sea horse Hippocampus erectus searobin Prionotus spp. silver hake Merluccius bitinearis margined snake eel Ophicthus cruentifer spot Leiostomus xanthurus striped anchovy Anchoa hepsetus weakfish Cynoscion regalis white hake Urophycis tenuis windowpane Scophthalmiis aquosus Invertebrates amphipods Cammarus spp. boreal squid Illex illecebrosus blue crab Callinectes sapidus cancer crab Cancer spp. channeled whelk Busycon canaticulatum fiddler crab Uca spp. lady crab Ovalipes ocellatus long-finned squid Loligo pealei mole crab Emerita talpoida sand shrimp Crangon spp. 1995 (Table 4). There were several commercially important species consumed by adult bluefish on Georges Bank in 1994. These included American pla- ice, haddock, Atlantic cod, and several hake species. A large component of the diet of adult bluefish from Cape Hatteras to Montauk Point, NY, was bay an- chovy. Butterfish, round herring, and squid were also important prey of bluefish collected from this geo- graphical region. The diet of bluefish from both re- gions also included several other fish and inverte- brate species (Table 4). Channeled whelk were preyed upon by adult bluefish in the same region and year that spring-spawned YOY bluefish were found to feed on them. Net feeding Although there was evidence of net feeding, it did not seem to bias diet indices. The percent occurrence of several different prey found to be "fresh" and the percent occurrence of these same prey found to be "digested" was similar (Table 5). Only in spring- spawned bluefish in the SNE region in 1995 were there more fresh prey than digested prey in their diet (30 fresh vs. 19 digested; Table 5). Spring- spawned bluefish had more freshly eaten prey in their diet (20—61%) than summer-spawned bluefish (0-267^). Prey-type selectivity Spring-spawned YOY bluefish selected positively icol/m) for bay anchovy in both 1994 (a=0.69, t-test vs. 0.25, t=8.05, df= 154, P<0.001 )) and 1995 ( a=0.80, ^test vs. 0.33, ^=6.40, df=74. P<0.001 ). However, YOY bluefish avoided (a CO CM 0 ^ CO oo -^ oi GO r-^ QJ ^1 <- ^ ^ QO UO CO CM 1— < CD CM CM CD Q CO O CO J3 a. c6 CO CM -^ l--^ '^ CD ^ 0 ^ CJi CO 0 CO 05 CO CO II a) t ° - 10 CO CM 0 CM 10 2 1 o6 »H CO "^ '^ O^ CM CO 00 5 CM o CO CO 00 in GO 0:1 oi crj CTi CD CO c^ ^ CO ^ ^ CO ^ ■* ^ CO -^ s ^ ■^ "^ 10 r^ .— 1 00 t— CD 0 CO 1^ ^ ITS u6 '^ 0 ■rr oi 1—1 oi s ^ " Ol ^ ^ CO uO 0 16 ^ 0 0 CD ^ CM crj 3 *J lO *J X 3 tio « -s ^ 0 o -^ . ^ Tj^ CO 0 I^ uo CO C O CO ■c " ^ CO >> a nal M le pre r sam CM ^ o ad iri, co CO CO QJ ^ o x^ a CM " .S CO O CO o CD o6 CO ^ § 00 CO CO CO c a ■^ CO CO CO CO Table 995 N identi for cli 00 LO 1 1 T CO 2 CO CO 0 "^ CD Oi CD 0 O Ql k. Tj^ UO CO CM 10 CM CM CM '"* -H ^ CO ■^ ° 3 o 5 UO 0:1 05 O iTS LO in C C M iri CM CM c; t>; 10 -<-> o ^ CO 0 ^ 0 0 0 0 CO '^ = ? 5 S '^ CO ^ 01 CM __, ■^^ oi 00 Oi CO ed dur portioi from ' Ol CO CO s oi m s CD 0 CM '^ GO CO CM CO CM oi 1 OS iQ fc. ^ t>^ ^ 16 CM CM :^ '"' CM o ^ I- r- CTi CO CD p 00 ptun =pro error LO ■^ CO ^ 0 CO ^ ^ (U «J CO > - Ol „ _ _ _ 0 CO ■^ ^ E spawned bluefish \g a prey (±SE), % 100%. SE=standar 0 CO CM CD CD 0 0 CO CS 13 CD tU) ^ c ^ w Z ^ ^ CO 06 CO 0 GO Oi CM 0 5 0 CO p 0^ CM 0 CM 0 V 0 CO CM QO 1^- CM QO 0 CM 0 r- CM 1 CO CO Oi CD m CM 1 uO CO g •s. 3 CO CO UO CM CO CD CO CM 0 CO 0 CO p CM CM CO tx.S o t- "-■ c3 c C -w >- a; C 'C 'S & TD 3 3 Q. -.J 3 Qj a bo C 'c5 'c r intents of s omachs con do not add c 'V. > 0 "3 (X '3 X Q. £ X at N CQ C X c/: £ E 3 a . 0. ^ to CO cc > 0 c Qi -T3 X -a 0 c 0 s 0 c c £ E b£ bJD il o X u c _;*; C 'br CO Qj S ^ 0 -0 t^ X ■0 c<3 CJ c TO c ho G3 Stomach C( bluefish st estimators 5> C 0 c C13 -a a 0 X u 0 X 7^ "c cfl ca a C3 a c 03 C 1-* bC i 0 a a c -0 CO c^ £ X £ c CO xT jf S 2 >> 22 -^ 3 J3 3 3 > X at CO 3 5 'S > ^ C3 0 3 3 0 0 3 2: 0 01 ^ -a -a Cu [JH H — M 764 Fishery Bulletin 97(4), 1999 <« t. O 01 -^ .2 = ■e " as !=§ II tn O CO fe ^ ID O CD 5 m CD r- ^ CO O 1 o O CD c^ 3 o at o c5 ID CD CO CO CO UO CO CM -^ i a UD % EC A » ^ # CO CO 1 £ CD co in GO CO CO o o in 1 o CO CT) in 1 CM O .— . I— 1 m ^ CO s CO O •^ w ^Q o c^i c^ IS CO CO 5 w iC ci 1:1 Tj" uo f^ o in ■^ :o 1^ a> -a ^ '* o 1—1 00 T3 si (U bo j= a eg 1 ^ k CO CO CO o OS -* CO in CO o o 3 CM CD O OO CM CO 1 c> ta _2 ■z ^ — — — 1— 1 01 2 CD o^ c— ir- C _ ,; CO c- '* -? o tn (M (N Tf t. -^ 01 ^ >.'H. O -* 00 CO o (N ^ UO TO M3 ^ lO OO CO CO C (D ^ C;^ '* .2 3 fe o cc # o o (M CO o^ CM CO o ^ Table 1995 N; identif for clu j= s. 3 « 2 M CO ^ tt, UO o o o O o UO o CZJ UO CM o o ira CD ^ CM i CO CM CO en 1 1^ ,"" o ^ c^ o 25 CM CM -5 = 0 ir- CO t^ tr- t- a c a cb CO CD CD cb ^ .2 S CO .— 1 : 199 ribut esti o ^ o l—t CD '^. 00 CD J3 c 2 1 ^ [r- CO CD t^ c r- CO ■" ° S OJ «, ^ c— OO rf r^ OO ^ O ho " ™ ■^ £ 1— i o CO o o o CO — b C > 05 t^ CM CD CD CM ■^ CO in -§■2 S 05 IS lf> OO ^ r- CO ,_^ CD oi ^ CD UO CO CO o CM o -a u p 1*. oi o CO co ■<^ CO o CM <" 9. i: c^ CO Cvl CO CM CO o ^ 3 O >- !>; ,_, QO CD "^ OS CD ^ bluefish capt ±SE), %W=pr standard erro CO CO 00 ■^ CO c^ CO CO CM ~ !>; t^ CO ^ t—' CO CO in El ^ o o ^ o o -^ o p o o a^ V V V V . — . o OS -a - II (U bo § in •"I CO o ■^ 2 >.W ■5 « CD in CO '-^ CD C g M 3 W 2 — _ CM CD CM CO 1*. o M Oi ■^ o o o LO 00 CO CO ci ai c^ c^ 'S- ■:^ o O a) c _ o CO CD CO CO lO ^ cri CD CJ e c 2 CO '~* V >. 6 -5 o- 01 7 c Qj 3 *i 3 a. c -o intents of s omachs con do not add JZ ^ c c « '« Qi w '3 — £ C o « > t cr cn S -a a* CO QJ '5 c o B B C cd 3 .i. Stomach ci bluefish St estimators ,- a c M CO xi c c o C 2 = -s <= ~ o a to -a to to O 't- x: CO £ o 1- -J C C3 C c J2 J3 < a s O c to o p. O to Id o II J fc 0^ *J 1:3 s ^ — ^l Bucket et al : Foraging habits of Pomatomus sa/tatnx 765 Table 4 Stomach contents of adult bluefish captured during the 1994 and 1995 National Marine Fisheries Service autumn bottom trawl survey. %F=proportion of bluefish stomachs containing a prey (±SE). %W=proportional contribution of identifiable prey to blue- fish diet by weight (±SE). Note that proportions from cluster estimators do not add up to 100*^. SE=standard error from variance estimators for cluster samples. 1994 1995 Georges Bank Hatteras-Montauk Georges Bank Hatteras-Montauk Prey type %F %W %F %W %F %W %F %W Fish bay anchovy butterfish round herring striped anchovy bluefish planehead filefish Northern puffer Northern pipefish searobin weakfish spot sand lance scup Atlantic herring Atlantic mackerel fawn cusk eel conger eel ocean pout fourbead reckling American plaice windowpane silver hake white hake red hake haddock Atlantic cod Gadidae Unidentified fish Invertebrates long-finned squid boreal squid lady crab cancer crab unidentified crab amphipods unidentified shrimp polychaete channeled whelk Total stomachs analyzed Number containing prey Mean FL (mm, SE) FL range (mm) Mean wt (g, SE) Wt range (g) 25.9(51.8) 21.4(47.7) 36.4(34.6) 34.9(44.9) 27.8(28.7) 18.6(23.6) 9.4(18.5) 12.7(30.0) 7.6(38.0) 16.3(80.2) 5.6(12.0) 9.0(20.8) 3.7(9.2) 2.2(5.6) 6.8(17.4) 26.5(87.9) 7.6(41.8) 10.8(66.4) 3.7(9.1) 3.6(11.1) 1.8(7.9) 1.7(7.7) 2.2(21.4) 5.2(51.5) 3.9(12.3) 0.5(2.3) 0.5(2.3) 0.9(4.6) 5.0(13.2) 0.9(4.7) 0.2(1.4) 0.8(4.0) 1.9(7.5) 0.6(2.5) 1.8(6.2) 2.1 (8.8) 2.2 (21.9) 6.5(65.3) 0.5(2.3) 1.1 (5.4) 2.2(7.8) 3.0(12.9) 0.5(2.3) 0.5(2.3) 0.1 (0.6) 0.9(4.7) 2.7)7.91 1.5(4.5) 0.5(2.3) 0.1 (0.5) 7.4(17.6) 11.3(23.7) 0.9 (4.2) 0.9(4.3) 10.9(54.7) 17.6(85.0) 0.5(2.3) 0.5(2.3) 0.5(2.3) 0.1 (0.7) <0.1 1.5(7.2) 0.9(4.3) 0.9(4.3) 2.2(22.9) 2.2(22.9) 0.5(2.3) 0.3(1.7) 1.9(7.5) 1.7 (7.0) 1.9(7.6) 0.8(3.5) 5.6(22.01 5.6(22.0) 2.2(21.9) 5.2(51.9) 1.9(7.6) 1.0(4.2) 4.3(39.5) 8.4(76.3) 1.9(7.5) 2.4(9.9) 1.9(6.7) 2.1 (7.6) 1,9(7.4) 1.7(6.7) 1.9(7.5) 2.0(8.0) 2.2 (21.9) 0.8(8.4) 3.7(10.3) 6.7(20.6) 1.9(7.6) 1.9(7.6) 29.6(32.1) 26.1 (43.11 30.4(70.7) 21.9(23.4) 29.6(25.6) 23.7(29.4) 11.6(22.1) 11.1(23.7) 8.4(19.4) 3.7(15.0) 1.2(4.7) 1.9(7.2) 0.6(2.5) 0.9(4.1) 0.9(4.3) 5.2(11.3) 15.2(53.6) 18.1(89.2) 7.7(11.8) 10.9(27.0) 0.8(3.5) 2.7(22.9) 0.4(3.7) 1.3(4.6) 0.1(0.4) 0.9 (4.3) 4.3(30.8) 2.0(19.1) 2.0(6.8) 4.9(16.7) 9.0(33.2) 13.0(41.9) 50 45 653(10) 420-780 3837 (151) 1209-6151 65 45 451(18) 310-780 1546(170) 368-5965 44 32 607(17) 380-750 3082(225) 692-6530 116 84 391(11) 310-730 991 (106) 319-4570 766 Fishery Bulletin 97(4), 1999 the prey to predator FL ratios for summer-spawned bluefish were higher in 1995 than they were in 1994 (Fig. 3). Summer-spawned bluefish consumed rela- tively larger bay anchovies in 1995 than in 1994; this was due to both the smaller size of the summer- spawned cohort in 1995 and the larger bay anchovy prey. A linear function describing bluefish capture success as a function of prey length to bluefish length ratio from Scharf et al. ( 1998b) is also plotted in Fig- ure 3 (capture success data for bluefish feeding on Atlantic silversides). The distribution of ratios for both spring- and summer-spawned bluefish in 1994 and spring-spawned bluefish in 1995 are values for which bluefish have relatively high capture success; Table 5 The total number of prey (n i found either fresh or digested (fresh=no sign of digestion; digested=prey skinless to be- ing only identifiable by skeleton or shape) in spring- and summer-spawned young-of-the-year bluefish stomachs and the percentage contribution of bay anchovy, striped an- chovy, butterfish, and squid for these categories. Bluefish were captured during 1994 and 1995 National Manne Fish- eries Service autumn bottom trawl sur\'eys. Bluefish were collected from three geographical locations of the Mid- Atlantic Bight continental shelf (SNE=Southern New En- gland. C-D=Chesapeake Bay to Delaware Bay, and SOC= South of Chesapeake Bay, after Munch 1997). Spring- spawned Summer- spawned Location Prey type No. No. No. No. fresh digested fresh digested (%) (%) (%) i%) 1994 SNE C-D 199.5 SNE C-D bay anchovy striped anchovy butterfish squid n bay anchovy striped anchovy butterfish squid bay anchovy striped anchovy butterfish squid n bay anchovy striped anchovy butterfish squid 19 89.5 5.3 5.3 0 46 89.1 2.2 2.2 6.5 30 73.3 6.7 3.3 16.7 21 81.0 19.0 0 0 51 56.9 7.8 33.3 2.0 107 76.6 2.8 7.5 13.1 19 68.4 10.5 15.8 5.3 82 82.8 13.8 0 3.4 36 100 0 0 0 31 104 97.1 0 1.0 2.0 57 36 160 140 120 100 H 80 60 40 20 D D »o D o<5b 2f ^*aiKi*^ 160 140 120 100 80 60 40 20 0 "1 r B 50 I I 1 1 — 100 150 200 250 Bluefish fork length (mm) 300 • bay anchovy o stnped anchovy D butterfish ▲ round hermng O sea robin larvae V squid Figure 1 Prey total length i mantle length for squid I versus YOY bluefish fork length in (A) 1994 {prey TL=0.134 x blue- fish FL+20.S2&. r2=0.10, P<0.0001) and (B) 1995 (prey TL=0.041 X bluefish FZ.-h44.538, r2=0.02, P=0.027). Table 6 Prey selectivity (Chesson's n, see text for calculations) in YOY bluefish collected on the U.S. east coast continental shelf in the autumn of 1994 and 1995. Values of a =l/m (where "m" is the number of prey categories) represent ran- dom feeding, whereas values of or > 1/m or a < 1/m repre- sent "selection" and "avoidance" of prey, respectively. Val- ues significantly different from 1/m (/-test, P<0.05) are in- dicated by a (-(-) for "selection" or I-) for "avoidance". Prey type 1994 (l/m=0.25) 1995 (l/m=0.33) Bay anchovy Butterfish Squid Other 0.69 (-h) 0.12 (-) 0.09 (-) on (-1 0.80 ( -I") 0.03 (-) 0.18 (-) Buckel et al.: Foraging habits of Pomatomus saltatnx 767 400 - A o 300 - ^o oo 200 - A9 **^ 100 - J i^ ^^ 0 - 1 1 1 II 1 ■ 400 - B o 300 - o o o 200 - r^ CP o A 0 3v^^ 100 - 0 - 1 1 1 ^|o 1 1 1 200 300 400 500 600 700 800 Bluefish fork length (mm) • bay anchovy n butterfish ▲ round herring o Atlantic hen-lng V squid ♦ crab o Other Figure 2 Prey total length ( mantle length for squid ) versus adult bluefish fork length in (A) 1994 iprey TL=0.199 ' blue- fish FL-14.413, r2=0.15. P<0.001) and (B) 1995 iprey TL=0.205 X bluefish FL-9.047. r2=0.28, P<0.0001). "Other" prey include searobin, red hake, silver hake, unidentified gadid, scup, windowpane, haddock. North- ern puffer, conger eel, cusk eel, sand lance, striped an- chovy, spot, weakfish, Atlantic mackerel, fourbeard rockling, and ocean pout. however, the prey length to predator length ratios of summer-spawned bluefish in 1995 are values at which bluefish have relatively lower capture success (Fig. 3). In 1994, there were four stations that had n>15 bay anchovy measurements for both spring and sum- mer-spawned bluefish (/i = 16-20 bay anchovy mea- surements for summer- and /! =22-39 for spring- spawned). Both spring- and summer-spawned YOY bluefish showed significant selection for relatively small bay anchovies (spring-spawned a=0.57 vs. 0.25, ^=2.99, df=38, P=0.005; summer-spawned a=0.65 vs. 0.50 ~[ 0.45 - 0.40 0.35 0.30 0.25 0.20 015 0.10 H 0.05 Spring-spawned Summer-spawned 1.0 08 h 0.6 0.4 0.2 'P^^r I — r- 0 1 0.2 0.3 0.4 0.5 0.6 0.7 0.8 0.1 02 0 3 0.4 0.5 0.6 07 08 Bay anchovy FL: bluefish FL ratio Figure 3 Bay anchovy FL to bluefish FL ratios for both sphng-and sum- mer-spawned bluefish in (Al 1994 and (B) 1995. Regression line for capture success versus prey length to predator length ratio from Scharfet al. (1998b). 0.33, ^=2.61, df=27, P=0.011) and avoidance of rela- tively larger anchovies in 1994 (spring-spawned a=0.02 vs. 0.25, ^=3.83, df=38, P<0.001: summer- spawned a=0.07 vs. 0.33, ;'=2.86, df=27, P=0.006) (Table 7). Intermediate-size bay anchovy were taken in proportion to their abundance in 1994 (^-test; P>0.05 for all or's). In the 1995 analysis, three sta- tions had n>15 individual bay anchovy lengths (« = 17-32); an analysis of size-selective feeding of summer-spawned fish in 1995 was not possible ow- ing to small sample size. Spring-spawned bluefish in 1995 showed no significant size selectivity (^test; P>0.05 for all or's); however, the trend of increasing selectivity with decreasing prey sizes was present (Table 7). 768 Fishery Bulletin 97(4), 1999 Feeding chronology, daily ration estimates, and Impacts on bay anchovy Both spring- and summer-spawned YOY bluefish gut fullness values varied over the diel cycle in the dif- ferent geographical regions of the MAB shelf. Peaks in gut fullness generally occurred at dawn, dusk, and diurnal time periods, whereas gut fullness values during nighttime collections were low in relation to diurnal gut fullness values (Fig. 4). Table 7 Size-selectivity (Chesson's a, see text for calculations) for bay anchovy in YOY bluefish collected on the U.S. east coast continental shelf in the autumn of 1994 and 1995. Values of o=l/m (where "m" is the number of prey catego- ries ) represent random feeding while values of a > 1/m or a < l/m represent "selection" and "avoidance" of prey, re- spectively. Values significantly different from l/m (?-test, P<0.05) are indicated by a ( + ) for "selection" or (-) for "avoidance". 1994 1995 Bay Spring- Summer- anchovy spawned spawned size (mm) (l/m=0.25) (l/ra=0.33) Spring- spawned Summer- ( l/m =0.33) spawned 25-34 35-44 45-54 55-64 0.57 (-t-) 0.27 0.14 0.02 (-1 0.65 (-H) 0.28 0.07 (-) 0.44 0.37 0.19 Table 8 Daily ration g/lg • d) • 100 ±SEl and weighted mean temperature at which spring- and summer-spawned YOY bluefish were collected during National Marine Fisher- ies Service autumn bottom trawl survey in 1994 and 1995. Bluefish juveniles were collected from three geographical areas of the Middle Atlantic Bight continental shelf iSNE=Southern New England, C-D=Chesapeake Bay to Delaware Bay, and SOC= South of Chesapeake Bay. after Munch 1997). Daily ration was calculated using the Eggers' ( 1979) approach (see text). NA=not applicable due to low number of samples for diel series. Summer-spawned Year and geographical region Temp. Daily ration g/(g-d)- 100 (SE) Temp. 1994 SNE C-D 1995 SNE C-D S( )( ' 200 18.9 61 20.7 8.5(3.1) NA 12.4(3.9) NA NA 182 129 79 177 51 18.8 20.0 19.9 23.7 24.3 In order to use the gut fullness data to estimate bluefish daily ration, gastric evacuation rate (GER) estimates were needed for bluefish at shelf water temperatures. The exponential GER model ad- equately described the evacuation of bay anchovy at 15 C from our laboratory experiment (/•-=0.87, n=22, P<0.001; Fig. 5). Incorporation of the 15°C GER es- timate (R^,-0.102) into the evacuation rate vs. tem- perature function of Buckel and Conover ( 1996) gave the following equation: fi^=0.017 e""«3^''""\ n = 16, r'^=0.82 (Fig. 5). The largest deviation from the equa- tion describing/?^, and temperature is from the single estimate of GER at 15"'C. This is likely not due to differences in experimental protocol; the 15°C GER experiment was conducted identically to the experi- ments described in Buckel and Conover (1996). The larger deviation may result from there being only one estimate of GER at 15°C whereas each GER estimate for the higher temperatures represents a mean of four estimates (Fig. 5). Estimates of GER from this equa- tion were used to estimate bluefish daily ration. The daily ration of summer-spawned bluefish in the SNE region in 1994 and 1995 was 8.5 and 12.4 g/ (g • d) ■ 100, respectively (Table 8). Spring-spawned bluefish daily ration estimates ranged from 4.8 to 6.6 g-/(g ■ d) • 100 for the SNE and C-D region in 1994 and 3.7-9.0 g/(g ■ d) • 100 in the SNE, C-D, and SOC regions of the MAB in 1995. There was insuffi- cient diel records to calculate feeding rates for the SOC region in 1994 or the remaining region and co- hort combinations (Table 8). The VPA estimates of YOY bluefish abundance in 1994 and 1995 were 24 million and 14 mil- lion, respectively. These estima- tions were partitioned into spring- and summer-spawned bluefish on the basis of relative abundances of these cohorts for each year (Table 9). It was esti- mated that spring-spawned bluefish consumed from 70 to 96 million bay anchovy per day in 1994 and from 68 to 170 mil- lion bay anchovy per day in 1995 during the September period of their migration (Table 9; the range is based on the lowest to highest daily ration estimate). Summer-spawned bluefish consumed 130 million bay anchovy per day in 1994 and 5 million per day in 1995 (there is no range for summer- spawned bluefish because there Spring-spawned Daily ration g/(g- d) ■ 100 (SE) 6.6(2.5) 4.8(1.9) 3.7(1.4) 9.0(2.9) 8.1 (2.8) Buckel et al,: Foraging habits of Pomatomus saltatnx 769 a was only one estimate of summer-spawned daily ration for each year). Discussion Dietary analyses Bay anchovy was the predominant fish prey of both spring- and summer-spawned YOY bluefish in both 1994 and 1995. A previous gut contents analysis of NEFSC- NMFS bottom-trawl-collected bluefish (south of Cape Hatteras to Georges Bank) was conducted by Morris^ from 1978 to 1980. Bay and striped anchovy dominated the diet of the 292 YOY bluefish (110- 300 mm) that he examined; butterfish and squid were also important prey. The mi- gration of YOY bluefish onto the shelf is thought to be controlled by declines in es- tuarine temperature and day length (011a and Studholme, 1972). Schools of bay an- chovy migrate out onto the shelf beginning in September and are concentrated off New York, New Jersey, and the Delmarva peninsula in the autumn (Vouglitois et al., 1987) in a similar pattern to that of con- centrations of YOY bluefish (Munch and Conover, in press). The timing of bluefish estuarine emigration may be linked with bay anchovy movements offshore. Invertebrates were an important part of the diet of YOY bluefish in several regions. In 1994, there was a large amount of gam- marid amphipods in the diets of spring- and summer-spawned bluefish in the SNE re- gion. In addition, mysids contributed to the diet of summer-spawned fish in the C-D and SOC regions of the continental shelf in 1994. Diets of juvenile bluefish in estuar- ies are usually dominated by fish prey, but at times invertebrates, such as shrimp, are a major prey (Friedland et al., 1988; Juanes and Conover, 1995). This pattern is likely a function of prey availability and prey size. Bluefish prey selectivity was related to prey abun- dance in the Hudson River (Buckel et al., 1999). Bay anchovy schools and invertebrate swarms are likely to have patchy distributions on the shelf; the diet of 0.00 008 v 3 0 06 0.04 an 0.02 0.00 en C 0.08 jP u.uo — Cl995-Summer-spawned A 0 06 - I /-\ 0.04 - 0 02 - 0 00 - D 1995-Spring-spawned \ Time intervals Figure 4 Gut fullness ig prey/g bluefish I for spring-and summer-spawned blue- fish collected from SNE, C-D. and SOC regions of the continental shelf in 1994 (A, Bl and 1995 iC, D) versus time interval of collection. Sun- rise fell between the 0600 and 0900 time inter\'al and sunset fell be- tween the 1800 and 2100 time interval. Error bars are ±1 SE. 2 Morris. T. L. 1984. Food of bluefish. Woods Hole Labora- tory Reference Document 84-22, Woods Hole Laboratory, North- Mar. Fish. Serv.', NOAA, 13 p. east Fisheries Science Center, Natl. Woods Hole. MA 02543 MED/NEFC 84-26. bluefish at any location will be dependent on local prey availability. Channeled whelk has not been reported previously as a prey of bluefish. The foot of channeled whelk with the operculum still attached was found in sev- eral bluefish in the C-D region in 1995. This may have been due to opportunistic feeding on whelk that had shell damage related to some fishery activity. However, whelk occurred in bluefish gut contents over a fairly broad geographic area and not at a single 770 Fishery Bulletin 97(4), 1999 station (Cape May, NJ, to Lewes, DE). Bluefish may attack whelk when the foot is extended out of the shell and may sever that section alone. This would explain the presence of only the foot portion of the animal. Summer-spawned bluefish diet was dominated by copepods in the SNE region in 1995. Marks and Conover ( 1993) found that the diet of summer-spawned bluefish collected in surface waters of the New York Bight was dominated by copepods. Summer-spawned bluefish size ranged from 17 to 64 mm TL in their samples and ranged from 20 to 140 mm FL in our samples from the SNE region in 1995. The large pres- ence of copepods in this cohort's diet compared with 1.0 -^ V y = 0 949 0-0^°^ ^'™ M •> • ■1 0.8 - »\ /•2 = 0.871 'n • ^ E P ^ • -. 0.6- \ • E '+-. ^B = 0.4 - o •\^ • 'S •^^ • a •^-J* S 0.2 A ^^~-^^ a. w 1 A n f\ - u.u 1 1 1 I 1 1 1 0 2 4 6 8 10 12 14 16 Time (h) 0.40 - 1 v= 0 017 e°°55 ''"'"P / 0.35 - / r2 = 0.818 W „ 0.30 - /^ 5 t/ 1 0.25- w 1 0.20 - / 3 g 0 15 ~ /^ u ^9 0.10 - • ^^ n ^^ B n n«; - U.UsJ 1 1 1 ! 1 1 ! 1 1 14 16 18 20 22 24 26 28 30 32 Temperature (X") Figure 5 (A) Proportion of meal remaining vs. time for blue- fish fed a single bay anchovy meal at 15-C {propor- tion of meal remaining = 0.949 e-o i»2 nm,.)_ and (B) YOY bluefish evacuation rate versus temperature 1 data for temperatures of 21 to 30'C are from Buckel and Conover, 1996;et'aci/a/;on rate=O.One'>"^^ "™p). Error bars are ±1 SE. other regions on the shelf in 1995 and with summer- spawned bluefish diets in 1994 is most likely a result of their small size (see "Prey-size analysis" section). McBride et al. (1995) found that age-0, spring- spawned bluefish abundance appeared to be regu- lated by density-dependent (compensatory) losses in Narragansett Bay, Rhode Island. One potential den- sity-dependent mechanism is cannibalism. There were very few cases of bluefish cannibalism on the continental shelf; spring-spawned bluefish cannibal- ism of summer-spawned bluefish occurred in the C-D and SOC region in 1995, a year when individuals in the summer-spawned cohort were relatively small in size. However, past diet studies of bluefish have found a higher incidence of cannibalism, particularly off the Carolinas (Lassiter, 1962; Naughton and Saloman, 1984). Bluefish abundance is currently low (NEFSCM and cannibalism may be frequent only when bluefish are more abundant or other prey abundances are low. We found slight regional differences in the diet of adult bluefish; diets of bluefish in 1994 and 1995 were similar to those found in past studies. In the Georges Bank region in 1994 and 1995, the dominant prey of adult bluefish were butterfish, long-finned squid, boreal squid, round herring, and Atlantic herring. Bay anchovy, butterfish, round herring, and long- finned squid were the dominant prey of adult blue- fish collected from Cape Hatteras, NC, to Montauk Point, NY. Morris (1984) found that long-fmned squid, boreal squid, butterfish, round herring, and bay and striped anchovies dominated the diet of adult bluefish (« =226) captured in NEFSC-NMFS bottom trawls ( 1978-80, spring, summer, autumn combined ) from these same regions. Richards ( 1976) examined the diet of adult bluefish angled near Long Island, NY, in the summer and autumn. The diet of bluefish in this area included bay anchovy, long-finned squid, menhaden iBrevoortia tyrannus), butterfish, and sil- ver hake (n=66, FL range: 490-750 mm). There were very few bluefish for our diet analysis from south of Cape Hatteras; however, Naughton and Saloman (1984) collected 729 bluefish from this re- gion using hook and line during 1977-81 and re- ported the diet of bluefish in this area as dominated by Scianidae, Clupeidae, Mugilidae, Labridae, and Atherinidae. Scianids, clupeids, engraulids, sparids, atherinids, and squids were important prey in Lassiter's ( 1962 ) study of over 900 bluefish captured by beach haul seining or hook and line August-Decem- ber 1960 and March, Jun^August 1961. The diet of bluefish (« =42) collected south of Cape Hatteras in the autumn by bottom trawling was dominated by squid, butterfish, and striped anchovy (Morris, 1984). There were several commercially important ground- fish species in the diet of adult bluefish on Georges Buckel et al.: Foraging habits of Pomatomus saltatnx 771 Table 9 Estimated spring- and summer-spawned bluefish daily consumption of bay anchovy on the U.S. east coast continental shelf in the autumn of 1994 and 1995. We used mean size of bluefish, daily ration estimates, proportion of bay anchovy in the diet, and mean bay anchovy size that were measured for each cohort in this study. Estimates of the numbers of YOY bluefish for 1994 and 1995 are from virtual population analysis (VPA) performed by NEFSC (see Footnote 1 in main text); we partitioned the numbers between spring- and summer- spawned bluefish using relative abundances of each cohort from the NEFSC-NMFS autumn ground- fish survey cruises. Cohort Biomass of Bluefish bluefish cohort numbers (10* kg) Amount of bay anchovy consumed daily (kg/d; low ration ) Amount of bay anchovy consumed daily (kg/d; high ration) Number of bay anchovy consumed daily (no./d; low) Number of bay anchovy consumed daily (no./d; high) 1994 Total 24 X \0^ Spring-spawned 7.2 X 10« 1.20 25,000 Summer-spawned 16.8 X 10« 0.50 35,000 1995 Total 14 X 10« Spring-spawned 9.8 X 106 1.86 45,000 Summer-spawned 4.2 V 106 0.032 2,000 34,000 110,000 70 X 106 130 X 106 68 X 106 5 X 106 96 X 106 170 X 106 Bank in 1994. Morris (1984) also found gadids in stomach contents of adult bluefish. This finding is of interest because there is a large effort to under- stand the factors that regulate fish populations on Georges Bank (Peterson and Powell, 1991). The bio- mass of skates (Rajidae) and dogfish sharks (Squalidae) has increased in this area of the shelf and in some locations makes up the largest part of fish biomass (Overholtz et al., 1991). However, carti- laginous fishes were not found in the diet of adult bluefish. Bluefish predation will have no direct ef- fect in regulating the population sizes of these fishes. However, bluefish may play a role in the recovery of commercially important groundfish species on Georges Bank. Because bluefish distribution is closely linked with temperature (Munch, 1997), an increased warming trend may allow a larger propor- tion of the bluefish population to extend northward onto Georges Bank. Ware and McFarlane (1995) found that Pacific hake (Merlucciiis productus) bio- mass and hake predation on herring (Clupea harengus) increased with recent increases in tem- perature off the west coast of Vancouver Island. They concluded that this recent increased predation ex- plains recent declines in the herring stock. The im- portance of bluefish predation to recovery of ground- fish species warrants further investigation. Net feeding Postcapture net-feeding can bias diet indices and gut fullness level estimates. Net feeding is known to oc- cur in a variety of Pacific midwater fish (Lancraft and Robison, 1980). There was little evidence that net feeding biased our estimates of bluefish diet given the similarities between percent of fresh versus di- gested prey; however, any net feeding would lead to biased estimates of gut fullness level and inflated estimates of consumption rate. Lancraft and Robison (1980) found that larger fish were more likely to in- gest artificial prey. The larger spring-spawned co- hort had more fresh prey in its diet than the smaller summer-spawned bluefish which may be a result of a higher incidence of net feeding by this cohort. Prey-type selectivity Spring-spawned YOY bluefish selected for bay an- chovy over all other prey tjrpes in both 1994 and 1995. Butterfish, squid, and "other" potential prey were avoided. This is most likely due to both the relative abundances of these prey as well as interspecific size differences. Bay anchovy tended to be the smallest prey available; the only smaller prey of spring- spawned bluefish were invertebrates such as amphi- pods (ranging from 3 to 15 mm TL). In the Hudson River estuary, the prey with the highest relative abundance were selected for, whereas the prey with the lowest abundance were selected against (Buckel et al., 1999). A second possible explanation for the strong selection of bay anchovy could be sam- pling bias. Chesson's ( 1978) index assumes that prey abundance is large in relation to the amount con- sumed (likely true for the shelf) and that the 772 Fishery Bulletin 97(4), 1999 catchability or availability of the different prey groups to the trawl are the same. We do not have estimates of the catchability for each of the prey spe- cies used in this analysis. Our results would be bi- ased if these differ. For example, catchabilities for squid and butterfish may be higher than that of bay anchovy and may explain the strong selection for bay anchovy. Prey-size analysis Although there was a weak positive linear relation- ship between ingested bluefish prey sizes and blue- fish length, it was mainly due to prey other than bay anchovy. The slope of the regression of bay anchovy length on bluefish length alone was 50% and 75% less than the slope of the regression of all prey on bluefish length in 1994 and 1995, respectively. The sizes of bay anchovy taken by the smallest YOY blue- fish were similar to those taken by the largest YOY bluefish, especially in 1995. Juanes and Conover ( 1995 ) and Scharf et al. ( 1997 ) saw a similar relation for blue- fish feeding on piscine prey in New York estuaries. Most ratios of bay anchovy length to bluefish length were similar to those seen in past estuarine work (Juanes and Conover, 1995; Scharf et al., 1997). Juanes and Conover (1995) hypothesized that the shift in spring-spawned bluefish diet from Atlantic silversides to bay anchovy in Great South Bay, NY, was a result of the higher capture success on the rela- tively smaller bay anchovy. The capture success of bluefish feeding on bay anchovy is likely higher than that of bluefish feeding on the relatively larger but- terfish and squid. Our size selectivity analysis shows that even within bay anchovy there is preference for smaller sizes (Table 7). The relatively small size of the 1995 SNE sum- mer-spawned bluefish (1994 mean FL=135 vs. 1995 mean FL=70) influenced the extent of piscivorous feeding in this cohort. The capture success was plot- ted for bluefish feeding at different ratios of prey length to bluefish length from laboratory work on Atlantic silversides (Scharf et al., 1998b). Summer- spawned, small-size bluefish in 1995 would have experienced greatly reduced capture success when feeding on bay anchovy. Feeding chronology, daily ration estimates, and impacts on bay anchovy Bluefish appear to feed mainly during the day or at dawn or dusk. This was also noted in past work on juvenile bluefish in estuarine environments (Juanes and Conover, 1994; Buckel and Conover, 1997). In the Hudson River estuary, the diel feeding patterns of bluefish appeared to be linked to their diel move- ment patterns ( Buckel and Conover, 1997 ). This may also be the case for shelf bluefish; Munch ( 1997 ) found that the catch per unit of effort (CPUE) ofYOY blue- fish in NEFSC-NMFS bottom trawl collections was significantly higher in diurnal than in nocturnal col- lections. Munch (1997) hypothesized that this diel movement to near bottom habitat (which makes them available to bottom trawling) during the day may be in response to the vertical migration of bay anchovy. Bay anchovy exhibit diel vertical migration and are significantly more available to bottom trawl sampling during the day compared with at night ( Vouglitois et al., 1987). Our daily ration estimates for spring-spawned bluefish (3.7-9.0 g Ag • d) • 100) are similar to past estimates of daily ration for September. Measure- ments of bluefish maximum daily ration from a labo- ratory mesocosm experiment (18.5-23°C) for mid- September were -10 g/(g ■ d) • 100 (Buckel et al., 1995). Field estimates of bluefish daily ration in the Hudson River estuary were 10.1-12.6 in mid to late August and 7.3 in mid-September (Buckel and Conover, 1997). Mean spring-spawned bluefish size was larger and mean temperatures were slightly lower on the shelf compared with bluefish sizes and temperatures in the above laboratory and field ex- periments. This finding may explain our slightly lower daily ration estimates for spring-spawned blue- fish on the shelf in the autumn. Hartman and Brandt's (1995b) laboratory estimates of maximum daily ration (C,,^^^) are also similar to our field esti- mates of spring-spawned bluefish daily ration esti- mates on the shelf From their experiments, esti- mates of C„|g, for bluefish ranging from 100 to 350 g at 20°C (temperature at which bluefish were found on the shelD are 7.5 to 12.5 g/(g • d) ■ 100. Owing to low sample sizes and an inadequate diel record, the daily ration of summer-spawned bluefish was determined only in the SNE region in 1994 and 1995. The daily ration estimate in 1994 was lower at 8.5 g/(g • d) • 100 than the estimate of 12.5 g/(g ■ d ) • 100 in 1995. The mean size offish in this cohort varied substantially between 1994 (31.2 g) and 1995 (4.8 g). The size and the temperature difference (1994= 18.9°C vs. 1995=20. 7°C ) may explain the differences in daily ration. The daily ration estimates of these fish are slightly lower than would be predicted from laboratory estimates of maximum daily ration of bluefish at these sizes and temperatures (Buckel et al., 1995; Hartman and Brandt, 1995b); however, both shelf estimates fall into the 95% confidence in- terval of the summer-spawned bluefish daily ration estimate (19-20 September 1992) in the Hudson River (Buckel and Conover, 1997). Buckel et al : Foraging habits of Pomatomus saltatrix 773 For our estimates of bay anchovy consumption by the YOY bluefish population during their Septem- ber migration, it was assumed that the entire YOY bluefish population occurred on the shelf ft'om Cape Hatteras, NC, to Montauk, NY. Support for this comes from sharp declines in YOY bluefish CPUE in early autumn estuarine beach seine surveys (Nyman and Conover, 1988; McBride and Conover, 1991) and an abrupt decline in average bluefish size on the shelf during early autumn (Munch, 1997). Our estimates show that given a 30-d migration period in Septem- ber, the population of YOY bluefish could consume from 6.0 to 6.8 billion bay anchovies in 1994 and from 2.2 to 5.3 billion in 1995. Summer-spawned bluefish consumed two orders of magnitude less bay anchovy per day in 1995 (5 million) compared with 1994 (130 million). This was partly due to the low abundance of this cohort in 1995 (4.2 million ) compared with that in 1994 ( 16.8 million). The amount of bay anchovy in the diet of summer-spawned bluefish (80'7f in 1994 vs. 60^^ in 1995 ) and the absolute size of bay anchovy prey taken by the two cohorts (0.27 g in 1994 vs. 0.54 g in 1995) explained the remainder. Interannual variation in the size structure of the summer-spawned cohort and their predominant piscine prey (bay anchovy) can have dramatic effects on the predatory impact of this bluefish cohort on bay anchovy populations. Although the importance of size-dependent processes in fresh- water fish predator-prey interactions is well de- scribed (Kerfoot and Sih, 1987; Ebenman and Persson, 1988), their importance for piscivore-prey interactions in marine systems is just beginning to be recognized (Juanes and Conover, 1995). There are no estimates of coastwide bay anchovy abundance for 1994 and 1995; hence, no direct ex- amination of the impact of this estimated bay an- chovy loss to the continental shelf bay anchovy popu- lation was made. There are bay anchovy biomass density estimates for estuaries on the east coast of the U.S. Vouglitois et al. (1987) estimated that the bay anchovy standing crop in Barnegat Bay, NJ, ranged from 830 to 4830 kg/km^. In Chesapeake Bay, Luo and Brandt's (1993) bay anchovy biomass den- sity estimate for September was -8000 kg/km-. We calculated the standing stock of bay anchovy on the shelf by multiplying the estimated ranges of estua- rine bay anchovy densities by the area ( km- ) in which bluefish were collected on the shelf This area (km-) estimate was calculated by summing all the NEFSC stratum areas in which YOY bluefish were captured in the autumn groundfish survey in 1994 and 1995. During the month of September, bluefish (spring- "and summer-spawned combined) consumption of bay anchovy could account from ~2 to 22% of bay anchovy standing stock in 1994 and from 1% to 24% of the bay anchovy standing stock in 1995. Bay anchovy are probably less dense on the shelf than in the estu- ary and our estimates of bluefish impact may be un- derestimates. However, the latest bluefish stock as- sessment (Mid-Atlantic Fishery Management Coun- cil') has found that bluefish abundance may be lower than that reported in the VPA ( NEFSC i). This could mean that our impact estimates are overestimates. At the upper range of the impact estimates, it is clear that the YOY bluefish population on the continental shelf consume a significant quantity of bay anchovy biomass. The effect of this loss on the population dy- namics of bay anchovy or the fish community on the continental shelf is unknown. Acknowledgments We thank T. Hurst, S. Munch, personnel of the North- east Fisheries Science Center of the National Ma- rine Fisheries Service, and the crew of the KV Alba- tross IV for aiding in the collection of bluefish used in this study. We also thank A. Green for assistance with stomach contents analysis. S. Morgan, T. Rotunno, J. Galbraith, and A. Matthews aided in prey identification. S. Munch, J. Galbraith, P. Kostovik, and R. Rountree were instrumental in obtaining data from the NMFS database. M. Terceiro kindly pro- vided bluefish stock assessment information. We thank R. Cerrato, R. Cowen, G. Lopez, and two anony- mous reviewers for critical reviews. This research was funded by the National Oceanic and Atmospheric Administration (NOAA) Coastal Ocean Program and by NOAA award nos. NA90AA-D-SG078 and NA46RG0090 to the Research Foundation of SUNY for the New York Sea Grant Institute. This work was prepared for publication while J.A.B. held a National Research Council-NOAA Research Associateship. Literature cited Azarovitz, T. R. 1981. A brief historical review of the Woods Hole labora- tory trawl survey time series. Can. Spec. Publ. Fish. Aquat. Sci. 58:62-67. Buckel, J. A., and D. O. Conover. 1 996. Gastric evacuation rates of piscivorous young-of-the- year bluefish. Trans. Am. Fish. Soc. 125:591-599. 1997. Movements, feeding periods, and daily ration of pis- civorous young-of-the-year bluefish iPomatomus saltatrix) in the Hudson River estuary. Fish. Bull. 95:665-679. ^ Mid-Atlantic Fishery Management Council. 1998. Amend- ment 1 to the bluefish fishery management plan, 340 p. [Avail- able from Mid-Atlantic Fishery Management Council. Room 2115, Federal Building, 300 South New Street, Dover, DE 19904-6790.] 774 Fishery Bulletin 97(4), 1999 Buckel, J. A., D. O. Conover, N. D. Steinberg, and K. A. McKown. 1999. Impact of age-0 bluefish iPomatomus saltatrix) preda- tion on age-0 fishes in the Hudson River estuary; evidence for density-dependent loss of juvenile striped bass. Can. J. Fish. Aquat. Sci. 56:275-287. Buckel, J. A., N. D. Steinberg, and D. O. Conover. 1995. Effects of temperature, salinity, and fish size on growth and consumption of juvenile bluefish. J. Fish Biol. 47:696-706. Chesson, J. V 1978. Measuring preference in selective predation. Ecology 59:211-215. 1983. The estimation and analysis of preference and its relationship to foraging models. Ecology 64:1297-1304. Chiarella, L. A., and D. O. Conover. 1990. Spawning season and first-year growth of adult blue- fish from the New York Bight. Trans. Am. Fish. Soc. 119:455-462. Cochran, W. G. 1977. Sampling techniques. Wiley, New York, NY, 428 p. Ebenman, B., and L. Persson. 1988. Size-structured populations: ecology and evolution. Springer- Verlag, Berlin, Germany, 284 p. Eggers, D. 1979. Comments on some recent methods for estimating food consumption by fish. J. Fish. Res. Board Can. 36: 1018-1019. Friedland, K. D., G. C. Garman, A. J. Bejda, A. L. Studholme, and B. Olla. 1988. Interannual variation in diet and condition in juve- nile bluefish during estuarine residenc.v. Trans. Am. Fish. Soc. 117:474-479. Hare, J. A., and R. K. Cowen. 1996. Transport mechanisms of larval and pelagic juvenile bluefish IPomatomus saltatrix) from South Atlantic Bight spawning grounds to Middle Atlantic Bight nursery habitats. Limnol. Oceanogr. 41:1264-1280. Hartman, K. J., and S. B. Brandt. 1995a. Trophic resource partitioning, diets, and growth of sympatric estuarine predators. Trans. Am. Fish. Soc. 124:520-537. 1995b. Comparative energetics and the development of bioenergetics models for sympatric estuarine piscivores. Can. .J. Fish. Aquat. Sci. 52:1647-1666. Juanes, F., and D.O. Conover. 1994. Rapid growth, high feeding rates, and early piscivory in young-of-the-year bluefish iPomatomus saltatrix). Can. J. Fish. Aquat. Sci. 51:1752-1761. 1995. Size-structured piscivory: advection and the linkage between predator and prey recruitment in young-of-the- year bluefish. Mar Ecol. Prog. Sen 128:287-304. Juanes, F., J. A. Hare, and A. G. Miskiewicz. 1996. Comparing early life history strategies oiPomatomus saltatrix: a global approach. Mar. Freshwater Res. 47:365-379. Kendall, A. W., Jr., and L. A. Walford. 1979. Sources and distribution of bluefish, Pomatomus saltatrix. larvae and juveniles off the east coast of the United States. Fish. Bull. 77:21.3-227. Kerfoot, W. C, and A. Sih. 1987. Predation: direct and indirect impacts on aquatic com- munities. Univ. Press of New England, Hanover, NH, 386 p. Lancraft, T. M., and B. H. Robison. 1980. Evidence of postcapture ingestion by nudwater fishes in trawl nets. Fish. Bull. 77:713-715. Lassiter, R. R. 1962. Life history aspects of the bluefish, Pomatomus saltatrix (Linnaeus), from the coast of North Carolina. M.S. thesis. North Carolina State College, Raleigh, NC, 103 p. Luo, J. L., and S. B. Brandt. 1993. Bay anchovy Anchoa mitchilli production and eon- sumption in mid-Chesapeake Bay based on a bioenerget- ics model and acoustic measures offish abundance. Mar. Ecol. Prog. Ser. 98:223-236. Marks, R. E., and D. O. Conover. 1993. Ontogenetic shift in the diet of young-of-year blue- fish Pomatomus saltatrix during the oceanic phase of the early life history Fish. Bull. 91:97-106. McBride, R. S., and D. O. Conover. 1991. Recruitment of young-of-the-year bluefish Pomato- mus saltatrix to the New York Bight: variation in abun- dance and growth of spring- and summer-spawned cohorts. Mar Ecol. Prog. Ser 78:205-216. McBride, R. S., M. D. Scherer, and J. C. Powell. 1995. Correlated variations in abundance, size, growth, and loss rates of age-0 bluefish in a southern New England estuary Trans. Am. Fish. Soc. 124:898-910. Munch, S. B. 1997. Recruitment dynamics of bluefish, Pomatomus salta- trix, on the continental shelf from Cape Fear to Cape Cod, 1973-1995. M.S. thesis, State University of New York at Stony Brook, NY, 127 p. Munch, S. B., and D. O. Conover. In press. Recruitment dynamics of bluefish iPomatomus saltatrix) from Cape Hatteras to Cape Cod, 1973- 1995, ICES -J. Mar Sci. Naughton, S.P., and C.H. Saloman. 1984. Food of bluefish iPomatomus saltatrix) from the U.S. south Atlantic and Gulf of Mexico. U.S. Dep. Commer, NOAA Tech. Memo. NMFS-SEFC-150, 37 p. Nyman, R. M., and D. O. Conover. 1988. The relation between spawning season and the re- cruitment of young-of-the-year bluefish, Pomatomus saltatrix. to New York. Fish. Bull. 86:237-250. Olla, B. L., and A. L. Studholme. 1972. Daily and seasonal rhythms of activity in the blue- fish iPomatomus saltatrix). In H. E. Winn and B. L. Olla (eds.l. Behavior of marine animals, p. 303-326. Plenum Publishing, New York, NY. Overholtz, W. J., S. A. Murawski, and K. L. Foster. 1991. Impact of predatory fish, marine mammals, and sea- birds on the pelagic fish ecosystem of the northeastern USA. ICES Mar Sci, Symp. 193:198-208. Peterson, C, and T. Powell. 1991. What IS GLOBEC' GLOBEC News 1:1-2. Richards, S. W. 1976. Age, growth and food of bluefish from east-central Long Island Sound from July through November 1975. Trans. Am. Fish. .Soc. 105:523-525. Scharf, F. S., J. A. Buckel, F. Juanes, and D. O. Conover. 1997. Estimating piscine prey size from partial remains: testing for shifts in foraging mode by juvenile bluefish. Environ. Biol. Fishes 49:377-388. Scharf, F. S., R. M. Yetter, A. M. Summers, and F. Juanes. 1998a. Enhancing diet analyses of piscivorous fishes in the Northwest Atlantic through identification and reconstruc- tion of original prey sizes from ingested remains. Fish. Bull. 96:575-588, Scharf, F. S., J. A. Bucket, F. Juanes, and D. O. Conover. 1998b. Predation by juvenile piscivorous bluefish iPoma- tomus saltatrix): the influence of prey to predator size ratio Buckel et al.: Foraging habits of Pomatomus saltatnx 775 and prey type on predator capture success and prey profita- bility Can. J. Fish. Aquat. Sci. 55:1695-1703. Seber, G. A. F. 1973. The estimation of animal abundance, and related parameters. Hafner Press, New York, NY, 506 p. Vouglitois, J. J., K. W. Able, R. J. Kurtz, and K. A. Tighe. 1987. Life history and population dynamics of the bay an- chovy in New Jersey. Trans. Am. Fish. Soc. 116:141-153. Ware, D. M., and G. A. McFarlane. 1995. Climate-induced changes in Pacific hake (Mertuccius produclus ) abundance and pelagic community interactions in the Vancouver Island upwelling system. In R. J, Beamish (ed.). Climate change and northern fish popula- tions, p. 509-521. Can. Spec. Publ. Fish. Aquat. Sci. 776 Abstract.-The bluefish, Pomatomus saltatrix, has long been considered a key predator on U.S. east coast fish species. Many of its prey species are also landed by humans, but no compari- son of prey biomass harvested by blue- fish versus fishermen has been at- tempted previously. We used data on growth, mortality, gross growth effi- ciency, and abundance to model the to- tal prey consumption rate by bluefish at-the population level. This estimate and previously published information on diet were used to calculate the bio- mass of individual resource species "harvested" by bluefish. The prey bio- mass consumed by bluefish annually along the U.S. Atlantic coast is equal to eight times the biomass of the blue- fish population. Bluefish consume a much higher biomass of squid and but- terfish than is currently harvested by commercial fisheries for these species. Bluefish consumption of Atlantic men- haden, however, was below the current fisheries landings for this species. For resource species that are shared with bluefish, our findings highlight the need for multi-species assessment and management. Mutual prey of fish and humans: a comparison of biomass consumed by bluefish, Pomatomus saltatrix, with that harvested by fisheries* Jeffrey A. Buckel Marine Sciences Research Center State University of New York Stony Brook, New York 11794-5000 Present address: James J. Howard Marine Sciences Laboratory National Marine Fisheries Service, NOAA 74 Magruder Road Highlands, New Jersey 07732 E-mail address jbuckelgsh nmfsgov Michael J. Fogarty Northeast Fisheries Science Center National Marine Fishenes Service. NOAA Woods Hole, Massachusetts 02543 David O. Conover Marine Sciences Research Center State University of New York Stony Brook, New York 11794-5000 Manuscript accepted 11 January 1999. Fish. Bull. 97:776-785 (1999). The bluefish, Pomatomus saltatrix, has long been regarded as one of the most voracious piscivores of the western North Atlantic. In 1873 the U.S. Commissioner of Fisheries, S. F. Baird, stated in a report to con- gress that: "... I am quite inclined to assign to the bluefish the very first position among the most inju- rious influences that have affected the supply of fishes ... a daily loss of 25 hundred million pounds . . . their trail is marked by fragments of fish and the stain of blood . . ." (Baird, 1873). Baird was concerned with the possibility that the high abundance of bluefish during that period was the cause of concomitant declines in fish landings. Bluefish abundance along the U.S. Atlantic coast has been known to fluctuate dramatically over the past two centuries (Bigelow and Schroeder, 1953). The impact of variations in bluefish abundance on their principal prey as well as on the community structure of the conti- nental shelf as a whole may be sub- stantial (Clepper, 1979; Kerfoot and Sih, 1987). If the major prey of blue- fish are also harvested by humans, then the potential fishery landings of such species may be influenced strongly by bluefish abundance. Moreover, species management through adjustment of the fishing effort will likely be ineffective if the amount of prey consumed by a key natural predator is substantially larger than the landings and ig- nored as a variable component of natural mortality. In this study, we estimate the to- tal prey biomass consumed annu- ally by a given biomass of bluefish using a model developed by Pauly ' Contribution 1136 of the Marine Sciences Research Center, State University of New- York, Stony Brook, New York 11794. Buckel et al.: A comparison of biomass harvested by Pomatomus saltatrix with that harvested by fisheries 777 (1986). Previously published data on diet and recent estimates of bluefish population size along the U.S. Atlantic coast were used to determine the "harvest" of prey by bluefish. We then compared "harvesting" of prey by bluefish to the biomass harvested by the fishery for the east coast as a whole. Prey consump- tion by age class was calculated to determine at which age the largest impact on prey occurs. We present these analyses not as management tools but merely as gross estimates of the probable magnitude of the bluefish harvest of fishery species so as to stimulate more detailed studies. Methods Estimates of population consumption rate The total U.S. east coast bluefish population con- sumption of a given prey was calculated from esti- mates of diet, the biomass of the bluefish popula- tion, and food consumption rate. Bluefish diet infor- mation was available from the literature for various seasons and regions along the east coast (Lassiter, 1962; Richards, 1976; Buckel et al.. 1999a: Morris^: Naughton and Saloman-). Bluefish biomass esti- mates were available for 1982 to 1995 from a virtual population analysis (VPA) performed by the North- east Fisheries Science Center, National Marine Fish- eries Service, Woods Hole, MA (NEFSC). However, estimates of bluefish consumption rate were avail- able only for young-of-the-year (YOY) (Juanes and Conover, 1994; Buckel et al., 1995; Buckel and Conover, 1997) or for age-0 to age-2 bluefish in Chesa- peake Bay (Hartman and Brandt, 1995a). For this reason, we used the approach described below. The quantity of food consumed annually by the bluefish population was estimated by using Pauly's (1986) age-structured consumption rate model. The model is based on a growth function (e.g. von Bertalanffy growth function, VBGF), the instanta- neous total mortality rate (Z), and the relationship of gross growth efficiency to weight. Pauly's (1986) model in simplified form is T {dw/dt)xN, 1 Morris, T. L. 1984. Food of bluefish. Woods Hole Ref. Doc. 84-22, 13 p. [Available from Woods Hole Laboratory, North- east Fisheries Science Center, Natl. Mar. Fish. Serv., NOAA, Woods Hole, MA 02543.] 2 Naughton, S. P., and C. H. Saloman. 1984. Food of bluefish {Pomatomus saltatrix) from the U.S. south Atlantic and Gulf of Mexico. U.S. Dep. Commer, NOAA Tech. Memo. NMFS-SEFC- 150, 37 p. 3 NEFSC (Northeast Fisheries Science Center). 1997. Report of the 23rd Northeast Regional Stock Assessment Workshop (23rd SAW) Stock Assessment Review Committee (SARC) con- sensus summary of assessments. Northeast Fisheries Sci. Cent. Ref. Doc. 97-05, 105 p. Q B 1 K,. dt \ W,xN,dt where Q / B is the food consumption per unit popula- tion biomass (see descriptions in Pauly ( 1986), Pauly and Palomares (1987), and Palomares and Pauly (1989)). The numerator of the model represents the food consumption of the population (Q), which is de- termined from the growth increment (dw /dt), num- bers offish at age t (N^, determined from an initial arbitrary number of recruits), and the growth effi- ciency (/lj) for a given weight at age t. Values are integrated over age classes from ^,. to ^,„^,j (age at re- cruitment to a maximum age). The denominator is the biomass term (B) which is the weight of individual fish at age / ( W,) multiplied by the number of fish ( A^, ) at age t. This value is integrated over all cohorts from t^ to <„Qj- An annual Q/B estimate represents the number of times the population consumes its own weight per year (Pauly, 1986). Pauly and Palomares ( 1987) have provided a BA- SIC progi'am to calculate Q/B from VBGF param- eters, a parameter from the relationship describing K^ as a function of weight, and an estimate of total mortality (Z). Growth for the Q/B model is based on the VBGF model expressed in weight form: W,=W,.x[l-e-' ■"-'"']' /here t = time in years; weight of bluefish at age t; asymptotic weight; growth coefficient; age at weight zero; and exponent of the length-weight relation- ship (i.e. W=aL^). W, = k = tn = For bluefish, Wilk's (1977) VBGF parameters were used where k = 0.226, t^ = -0.123, W. = 8725g, and b = 3 (Wilk's estimate of 6 was 2.89 but 6 = 3 was used owing to ease of use in Pauly's ( 1986) model). The Q/B model assumes that gross growth effi- ciency (ifj ) data depends on weight ( W) according to A', = l-(W/W )/* To estimate (5 for use in the calculation of Q/B, blue- fish gross growth efficiency and weight data from experiments on YOY fish described in Buckel et al. (1995) were used. The linearized form of the above model described in Pauly ( 1986 ) was fitted to the data 778 Fishery Bulletin 97(4), 1999 to estimate i3(/3 =-0.0445, r2=0.64, F=45.44, df=l,25, P<0.0001). The current stock assessment of bluefish assumes a natural mortality of M=0.25. The highest fishing mortality (F) measured on the east coast bluefish population was in 1992 at 0.51 and recent estimates of Fare around 0.4-0.5 (NEFSC^). Therefore, an es- timate of M + i^=0.75 was used as a total mortality (Z) input into the Pauly ( 1986) model to estimate Q/B. The value of Z does not affect the estimate of Q/B dramatically because of its presence in both the nu- merator and denominator of the equation (see above). Annual consumption (Q/B) was estimated for blue- fish from age-0 to age- 10. To determine if our estimate was robust, Hartman and Brandt's ( 1995a) equation iC^^^ = 0.520 x W-o -88) describing laboratory (20-25°C) estimates of maxi- mum consumption rate iC^^^, g/(g • d)) as a function of bluefish weight (W, g) was also used to calculate the population consumption for 1995. First, bluefish biomass by age class was calculated by multiplying bluefish numbers at age (NEFSC^) by the mean blue- fish weight at age ( W, calculated from Wilk's ( 1977 1 VBGF function and length:weight conversion equa- tion; W^,^,^, Q was estimated at ^=0.5 ). Second, the maxi- mum consumption rate was calculated for a mean bluefish weight at age from the relationship between ^max ^"^ bluefish weight provided by Hartman and Brandt ( 1995a). To do so, we extrapolated out beyond the range of fish sizes used in their laboratory ex- periment, assuming that values obtained in this way were accurate. Lastly, these age-specific consump- tion rates were then multiplied by the biomass of each age class. Summing across age classes provided us with a maximum biomass of prey consumed by the bluefish population at 20-25°C; this value was divided by the biomass of the population to obtain a daily Q/B value. To calculate an annual Q/B value, the daily Q/B value was multiplied by 365 days. Additionally, this Q/B value was temperature ad- justed from 22.5°C to 17.5°C. This was done by mul- tiplying by the ratio of bluefish consumption rate estimated at 17.5°C (average temperature for coastal bluefish [Munch, 1997] ) to consumption rate esti- mated at 22.5°C (temperature of Hartman and Brandt's [1995a] study). This ratio was estimated from both Buckel et al. (1995) and Hartman and Brandt ( 1995a) data and in both cases was found to be -0.60. Impact by age class From the above analysis, the age class where total biomass consumption peaks can be determined. This calculation was first made for a simulated bluefish population having constant recruitment (N=l x IC) and total mortality (Z=0.75, see above). However, bluefish exhibit highly variable recruitment across years. To illustrate the impact of variable age struc- ture, we also calculated biomass consumed by age class on the basis of actual abundances in 1984 (a year when preceding recruitments had been fairly constant and age structure was typical) and in 1995 (a year when preceding recruitments were highly variable and age structure was unstable). Comparison of biomass consumed by bluefish with that harvested by fisheries The estimate of the annual amount of prey consumed by the western Atlantic bluefish population is the annual Q/B estimate (described above) multiplied by stock biomass. The prey consumption for three different population biomass sizes was estimated based on VPA estimates; these were the minimum, maximum, and average population sizes from 1982 to 1995 (NEFSC^). The total annual consumption at these three stock sizes was estimated for Atlantic butterfish (Pepriliis triacanthits), long-finned squid (Loligo pealei). boreal squid (Illex iUecebrosus). and Atlantic menhaden (Brevoortia tyrannus). These are primary species in the diet of bluefish that are also landed by fishermen. Several studies were used to attribute the total biomass consumption by bluefish to different geo- graphic regions, seasons, and prey species. To simu- late the migratory patterns of bluefish, it was as- sumed that 100% of the population was north of Cape Hatteras during summer and autumn and 100% of the population was south of Cape Hatteras during spring and winter. Diet data from Richards (1976), Morris^ and Buckel et al. (1999b) were used to de- scribe bluefish diet in the summer and autumn north of Cape Hatteras. The diet studies of Naughton and Saloman- and Lassiter (1962) were used to describe bluefish diet south of Cape Hatteras (Carolinas and southeast Florida) during winter and spring. Details regarding these diet studies can be found in Buckel et al. (1999b I. To calibrate the relative importance of bluefish predation on squid, butterfish, and menhaden, the annual bluefish consumption of these prey was plot- ted along with the average, minimum, and maximum fisheries landings ( 1984-92 ) for these species as com- piled by the National Marine Fisheries Service (Anonymous''). ■•Anonymous. 1993. Fisheries ofthe United States, 1992. U.S. Dep Commer., National Oceanic and Atmospheric Association. National Marine Fisheries Service. Silver Spring, MD, 115 p. Buckel et a\: A comparison of blomass harvested by Pomatomus saltatnx with that haroested by fisheries 779 Table 1 Estimate of daily and annual prey consumption of the 1995 bluefish population by Wilk,' maximum consumption rate estimates from Hartman and Brandt (1995a), estimates of bluefish number by age from NEFSC (see Footnote 3 in main text). age class. We used weight-at-age data from and the virtual population analysis (VPA) Age Bluefish weight (kg) Bluefish numbers (millions) Bluefish biomass ( 10*^ kg) Consumption rate (kgAkg- d)) Biomass of prey consumed daily (kg) 0.5 0.031 13.89 0.43 0.1941 82538 1 0.143 17.47 2.50 0.1245 311317 2 0.662 4.78 3.16 0.0801 253240 3 1.502 3.95 5.92 0.0633 374750 4 2.524 2.86 7.22 0.0545 393250 5 3.604 3.31 11.92 0.0492 585820 6 4.654 5.88 27. .35 0.0456 1249322 7 5.622 2.76 15.53 0.0432 671889 8 6.481 1.55 10.03 0.04152 416561 9+ 7.225 3.99 28.84 0.0402 1160642 Sum of bluefish biomass (B; 10*^ kg) 112.90 Sum of biomass consumed (Q; kg) 5499328 Q/B daily (at 22.5°C) 0.0487 Q/Bannual(at22.5''C) 17.78 Q/B annual (at 17.5°C) 10.84 ' Wilk, S. J. 1977. Biological and fisheries data on bluefish J. Howard Marine Sciences Laboratory, Northeast Fisheries Pomatomus saltatrix (Linnaeus Science Center, Natl. Mar. Fish- . Tech. Ser Rep. 11, 56 p. [Available from James Serv., NOAA. 74 Magruder Rd., Highlands. NJl Effects of pooling diet across ages One potential shortcoming of this analysis was the use of diet data pooled across age or size groups. This procedure was unavoidable because of the lack of data describing bluefish diet by age or size for different seasons and geographical locations on the east coast of the United States. The use of pooled diet data may bias impact estimates given that prey such as squid, butterfish, and squid are more important diet items of older and larger bluefish. To address this poten- tial bias, we performed a separate analysis to calcu- late bluefish consumption of prey using size or spe- cific diet data (where available, see Morris') as op- posed to pooled data. The daily biomass of prey con- sumed by each age or size class ( uncorrected for tem- perature) was calculated for the simulated popula- tion (a population with constant recruitment and mortality, see "Estimates of population consumption rate" section) and the 1995 bluefish population (a population whose biomass was dominated by older fish, see "Impact by age class" section and Table 1). These consumption-by-age estimates were multiplied by percent-contribution-by-weight data from Morris' for the four prey species described in the main analy- sis. The biomass of each prey type that was consumed was summed over all age classes to obtain the popu- lation consumption of each prey type. These values were then compared with values of population con- sumption of each prey type that were obtained by multiplying the total population consumption of all prey by the pooled prey contribution from Morris. Results Estimates of population consumption rate The annual QIB estimate from Pauly's (1986) model for the east coast bluefish population is 7.7. This means that the east coast bluefish population con- sumes a biomass of prey that is equivalent to ~8 times its own biomass over a one year period. The estimate of Q/B with Hartman and Brandt's ( 1995a) consump- tion-rate-at-weight data, adjusted for temperature, and estimated for the 1995 bluefish biomass on the U.S. east coast, was 10.8 (Table 1). Impact by age class For the simulated population with constant recruit- ment and mortality, the peak predatory impact of the bluefish population occurs between approxi- mately age 1 and age 3 (Fig. 1). In 1984, a year with 780 Fishery Bulletin 97(4), 1999 fairly typical age structure, peak absolute consump- tion of prey occurred between age 2 and age 3 with a second peak at age 6 (Fig. 2). In 1995, when age struc- ture was dominated by the 1989 year class, peak predatory impact occurred in fish that were six years of age (Table 1; Fig. 3). Therefore, much interannual variation in the age classes at which the peak preda- tory impact occurs should be expected in bluefish. Comparison of biomass consumed by bluefish with that harvested by fisheries The bluefish VPA biomass estimates from 1982 to 1995 ranged from a low of 124,024 metric tons (t) in 1995 to a high of 328,864 t in 1986 with an average stock size of 233,309 1 for this period (NEFSC^). Given a Q/B estimate of 7.7, the minimum-size bluefish 1e+7 - r ^ 1e+7 - A - 6 n ;; 8e+6 - \ ^ % E 6e+6 - \ ^ - 4 " 3 \ , ' ?S z 4e+6 - \^ • - 2 ^ 2e+6 - ....■^<^_^ - 0 Oe+0 ~~ 2e+6 - 1 1 1 1 « 1e+6 - B ^^^ _:.; 1 N. :; ie+6 - / \ 1 9e+5 - o / \ ^ 6e+5 - / \^ 3e+5 - / \^^ fio+n — 0 20 - 1 1 1 1 2 015 - \ c — \ O -3 \. 1^ 0 10- \v 5 5- ^^^^^^ § 0 05 - ' — ■ — — o 1e+5 - 1 1 1 1 ' ^ 9e+4 - / \ g_ 6e+4 - / \ 3e+4 - ^^^^ Ci^+Cl ^"^^ ue+u n 1 1 1 1 0 2 4 6 8 Age (years) Figure 1 Estimates of bluefish impact on prey by age class for a simu- lated population of the U.S. east coast bluefish stock. Simu- lated numbers of bluefish (see text) and bluefish weight (kg; Wilk,1977) at age (A) were used to calculate bluefish bio- mass (Bi. Impact (kg) on prey by each age class (D) was calculated by multiplying bluefish biomass by maximum con- sumption rate iHartman and Brandt, 1995a) for each age class (C). 1e+8 -[ q r ® 8e+7 - 1 6e+7 - 1 4e+7 - 2e+7 - - 4 do' - 9 ^ -'■ ■ - 0 4e+7 - J," 3e-^7 - 1 1 1 1 B /v /\ ; 3e+7 - / \ i 2e+7 - / ^-^ ^ 1e+7 - / 7e+6 - J n A ' 0 20 - 1 1 1 1 \C 1 0.15 - \ |I 0 10- V 7 ■-'' 5 " 0 05 - ^ O 2e+6 - 1 1 1 1 D /\ ^ 1e+6 - / ^^\ |- le+6 - 5e+5 - C 1 1 1 1 ) 2 4 6 8 Age (years) Figure 2 Estimates of bluefish impact on prey by age class for the U.S. east coast bluefish stock in 1984. Numbers of bluefish (NEFSC<) and bluefish weight (kg: Wilk, 1977) at age (A) were used to calculate bluefish biomass (Bl. Impact (kg) on prey by each age class (D) was calculated by multiplying bluefish biomass by the maximum consumption rate (Hartman and Brandt, 1995a) for each age class (C). Buckel et al,: A comparison of biomass harvested by Pomatomus saltatrix with that harvested by fisheries 781 population (1995) consumed 0.9 million t per year whereas the maximum-size population (1986) con- sumed 2.5 million t of prey per year. The 1982 to 1995 average population biomass consumed 1.8 mil- lion t of prey per year. Of the total biomass of prey consumed annually by bluefish, the fraction attributed to the primary resource species is an annual average of 10% butter- Oe+0 020 c ^ 0,15 I" 00 0,10 3 " ' 60 C o U 0 05 - 1e+6 - M 6e+5 - 3e+5 Oe+0 "I 1 r 2 4 6 Age (years) Figure 3 Estimates of bluefish impact on prey by age class for the U.S. east coast bluefish stock in 1995. Numbers of bluefish (NEFSC3) and bluefish weight (kg; Wilk. 1977) at age (A) were used to calculate bluefish biomass IB). Impact I kg l on prey by each age class (D) was calculated by multiplying bluefish biomass by maximum consumption rate (Hartman and Brandt 1995a I for each age class (C). fish, 10% long-fmned squid, 5% boreal squid, and 5% Atlantic menhaden. These are also averages across bluefish body size; however, these averages may be too simplistic given that bluefish diet changes with increasing size. Much of the rest of the diet of blue- fish is attributed to bay anchovy (described in Buckel et al., 1999a) for which no fishery landings exist. The average (minimum-maximum) annual consumption of butterfish by bluefish is therefore 90,000 1 (48,000- 125,000 t); annual consumtion of long-finned squid, is 90,000 t (48,000-125,000 t); and annual consump- tion of boreal squid, is 45,000 t (24,000-63,000 t). The annual consumption of these species by bluefish is much higher than the annual fisheries landings over the period 1984-92 (Fig. 4). The average con- sumption of Atlantic menhaden by bluefish was 45,000 t (24,000-63,000 t). The menhaden fishery takes seven times the biomass of menhaden that bluefish consume (Fig. 4). Effects of pooling diet across ages For the simulated population, the daily consumption of long-finned squid and butterfish calculated by using age-specific data did not differ greatly from daily consumption when diets were pooled across ages (Fig. 5). However, the use of pooled diets over- estimated daily consumption of boreal squid and At- lantic menhaden in comparison to values obtained with age-specific diets for both the simulated and 1995 bluefish population. For the 1995 population, age-specific diets gave daily consumption values for long-finned squid and butterfish that were higher than those that were calculated when pooled percent diet contributions were used (Fig. 5). Therefore, pool- ing diets can lead to under- or over-estimates of the predation pressure by bluefish; this bias can be affected by the age structure of the bluefish population. Discussion The results suggest that for at least three of the prey species of bluefish and humans, bluefish harvest a much higher biomass than does the fishery. This predatory loss will vary with fluctuations in blue- fish abundance. We elaborate on management im- plications below. Estimates of population consumption rate Our estimate of QIB (annual population consump- tion) is likely robust for two reasons. First, the esti- mate of bluefish annual QIB is similar to that of other pelagic piscivores. Palomares and Pauly (1989) esti- 782 Fishery Bulletin 97(4), 1999 150 -I g 100 75 - 50 25 Bluefish consumption Landings 400 mi - 200 2 boreal iong-fmned butlert'ish menhaden Figure 4 Estimated biomass < in thousands of metric tons ( t ) I of boreal squid, long-finned squid, butterfish. and Atlantic menhaden consumed by the U.S. east coast bluefish population (open histogram rep- resents consumption by average bluefish biomass from 1982 to 1995, error bar represents range for consumption by a minimum to maximum bluefish population size) and harvested by east coast fisheries (closed histogram represents average harvest from 1984 to 1992, error bar represents range for harvest from minimum to maximum I. mated the annual Q/B of bar jack, Caranx ruber, and the dolphin Coryphaena hippuriis at 10.6 and 8.5, respectively. Secondly, the estimate of Q/B with labo- ratory-measured daily rations (age 0—2 yr fish) and 1995 VPA bluefish biomass data gave us an estimate of Q/B of 10.8. This value was only slightly higher than the estimate from Pauly's (1986) model. This difference is likely a result of extrapolating C,„^,^ data from Hartman and Brandt {1995a) to larger fish. Assuming consumption rates of these larger fish were overestimated, values of 2.0 to 3.0 kgAkg ■ d) ■ 100 were substituted for fish ages 4-9-I-. From this a Q/B value of 6.6 was estimated. This value is more similar to the Pauly model Q/B estimate of 7.7. Future esti- mates of bluefish population consumption should include laboratory or field estimates of feeding rates of older bluefish. The Q/B estimate based on the 1995 VPA (10.8) incorporates the age structure existing at that time (recruitment not constant in the population), whereas the estimate from Pauly's (1986) model (7.7) assumes constant recruitment. Age structure may also con- tribute to the differences between the two Q/B esti- mates. For the 1995 bluefish stock, recent years of low recruitment have led to peak biomass in the older age classes (NEFSC*). These older, more abundant age classes consume a higher biomass of prey than younger age classes. In 1995, age-6 fish consumed more prey biomass than any other age class. This A Simulated boreal long-finned butterfish menhaden Figure 5 Daily consumption (not temperature corrected; t) of boreal squid, long-finned squid, butterfish, and At- lantic menhaden by the U.S. east coast bluefish population based on simulated (A) and 1995 (B) data ( see text for calculation methods ). Consumption was estimated from pooled diet indices (open histogram) and indices which were partitioned by age or size (closed histogram). strong age class resulted from a relatively large re- cruitment in 1989. If recruitment and mortality are constant, our analyses predict that absolute prey consumption will peak at about age 2 ( Fig. ID ). That consumption rate tends to peak in the earlier age classes has also been found in other species (Stewart and Binkowski, 1986; Yafiez-Arancibia et al., 1993). Bluefish ingest larger and a greater diversity of prey and include more resource species in their diet (e.g. butterfish, squid, Atlantic herring, and several groundfish species), with increasing body size (Buckel et al., 1999b; Morris'). When coupled with temporal shifts in the biomass age structure of bluefish, such ontogenetic changes in diet have important implica- tions for impact on prey populations. For example, bluefish recruitment was relatively high from 1982 through 1984. In 1984, peak absolute consumption of prey occurred between age 2 and 3 compared with age 6 in 1995. In 1984, the biomass consumption of age-0 through age-3 fish was 40% of the entire Buckel et al.; A comparison of biomass harvested by Pomatomus saltatrix with that harvested by fisheries 783 population's consumption, but it was only IS'yf of to- tal population consumption in 1995 (Figs. 2 and 3). Similarly, Hartman and Brandt ( 1995b) found that rela- tive prey consumption by bluefish at the population level in Chesapeake Bay varied with age structure. The fact that age structure and ontogenetic diet shifts interact provides further justification for ob- taining bluefish diet information by age or size. For example, the impact of the 1995 bluefish population on long-finned squid and butterfish was underesti- mated when pooled diet indices were used compared with diet indices that were partitioned by age or size (Fig. 5). Comparison of biomass consumed by bluefish with that harvested by fisheries Edwards and Bowman (1979) and Sissenwine et al. (1984) found that total piscivore consumption often exceeded fishery landings on Georges Bank. We have shown that bluefish alone consume an amount of squid and butterfish far exceeding the harvest of these species. How do our findings relate to the man- agement of squid and butterfish? The current man- agement plan for long-finned squid recommends a target yield of 21,000 t (Mid- Atlantic Fishery Man- agement Council^). From calculations presented here bluefish consumed almost five times this amount of long-finned squid. Consistent with our findings, es- timates of natural mortality for butterfish, long- finned squid, and boreal squid are high compared with other species (Anonymous, 1995; Atlantic but- terfish, M=0.80; long-finned squid, monthly M=0. 34; boreal squid, M>1.0). Our analysis does not allow us to determine what fraction of total prey mortality is due to bluefish consumption. Nor do we have esti- mates of impact of other predators on these prey. If we knew that bluefish were the dominant contribu- tor to natural mortality in these prey, then questions regarding allocation could be dealt with more explic- itly. For example, if the goal were to build larger squid stocks, would increased fishing mortality on blue- fish be more or less effective than reductions in fish- ing mortality on squid? Future research should deter- mine the magnitude of predation mortality resulting ft"om bluefish and other predators. Until this informa- tion is available we must assume that predation mor- tality is similar to or below estimates of natural mor- 5 Mid-Atlantic Fisheries Management Council. 1996. Amend- ment 6 to the fishery management plan and the draft environ- mental assessment for the Atlantic mackerel, squid, and but- terfish fisheries. [Available from Mid-Atlantic Fishery Man- agement Council, Room 2115, Federal Building, 300 South New Street, Dover, DE 19904-16790, 16 p.] tality for squid and buttei-fish. The management plan for these prey species assumes a high natural mortal- ity (high predation mortality already taken into ac- count) and fishing harvests are targeted accordingly. However, these stock assessments assume that natural mortality is a fixed value. For prey of blue- fish, this may not be true given the persistent changes in bluefish abundance on multiyear to decadal time scales. Natural mortality in squid and butterfish may vary greatly as a function of bluefish abundance. This has important management implications. For ex- ample, if bluefish abundance is high and remains so for extended periods and if this results in natural mortality rates on the prey that exceed the baseline natural mortality rates, then the biological reference points used in squid and butterfish management must be adjusted accordingly. Even if fishery remov- als are low in relation to those for other species, they may still contribute to population collapse by driv- ing the prey population past the replacement level. Much more attention needs to be focused on assess- ment of the dynamics and predatory impact of blue- fish if we wish to manage its prey, not to mention bluefish. Our results illustrate the importance of in- corporating interspecific interactions in the manage- ment of marine fisheries (Sissenwine and Dann, 1991; Magnusson, 1995). Bluefish may have limited influence on menhaden population dynamics because bluefish consumption of Atlantic menhaden was substantially below the fisheries landings of this species in our calculations. Oviatt (1977) estimated the aggregate demand for menhaden by bluefish in Narragansett Bay. She found that menhaden abundance was sufficient to meet the demands of bluefish even when the men- haden stock was low. Hartman and Brandt (1995b) found that combined predation by bluefish, weakfish, and striped bass on menhaden in Chesapeake Bay was low compared with fisheries harvest of menhaden. There are several limitations to this analysis. The estimates of the biomass of prey consumed rely on several other estimated parameters. These include the VPA of bluefish biomass, the allometry of gross growth efficiency as a function of bluefish weight, diet estimates, and assumptions about the spatial and temporal location of the bluefish population. We therefore urge readers to view our estimates as rough approximations of the true values. At the very least, however, our analysis clearly suggests a need to greatly improve our knowledge of the population dynamics and foraging ecology of bluefish. In par- ticular, more accurate data on bluefish abundance, spatial distribution, and diet as a function of blue- fish size, season of the year, and habitat (e.g. estu- ary, coastal shoreline, or offshore) are required. 784 Fishery Bulletin 97(4), 1999 For example, it has been shown that poohng blue- fish diet by age or size class may over- or under-esti- mate the impact of bluefish compared with estimates of impact obtained when using age or size partitioned diets. Additionally, the diet of bluefish in estuaries (Oviatt, 1977; Hartman and Brandt, 1995c; Buckel et al., 1999a), where YOY bluefish are known to be dominant piscivores, was not considered here. Buckel et al. (1999a) showed that YOY bluefish in the Hudson River estuary are key predators on YOY striped bass and that recruitment success of striped bass is negatively correlated with YOY bluefish abun- dance. In the Chesapeake Bay, the percentage of menhaden in bluefish diet can be over 807f of the diet by weight (Hartman and Brandt, 1995c), much higher than the diet proportion we used in our analy- ses. A synoptic study examining the diet on the shelf and estuarine environments, as well as temporal and spatial distributions, will be required to determine the exact proportions of prey items in the coastal bluefish population. Acknowledgments We thank M. Terceiro who kindly provided bluefish stock assessment information. Critical reviews were provided by R. Cerrato, R. Cowen, G. Lopez, and an anonymous reviewer. This research was funded by the National Oceanic and Atmospheric Administra- tion (NOAA) Coastal Ocean Program and by NOAA award no. NA90AA-D-SG078 and no. NA46RG0090 to the Research Foundation of SUNY for the New York Sea Grant Institute. This work was prepared for publication while J. A. B. held a National Research Council-NOAA Research Associateship. Literature cited Anonymous. 1995. Status of tlie fisliery resources off the northeastern United States for 1994. U.S. Dep. Commer, NOAA Tech. Memo. NMFS-NE.108. 140 p. Baird, S. F. 1873. Natural history of some of the more important food fishes of the south shore of New England. Part II: the bluefish. U.S. Commissioner of Fish and Fisheries, Wash- ington, D.C., p. 235-252. Bigelow, H. B., and W. C. Schroeder. 1953. Fishes of the Gulf of Maine. Fishery Bulletin of the Fish and Wildlife Service, Washington, D.C., 577 p. Buckel, J. A., and D. O. Conover. 1997. Movements, feeding chronology, and daily ration of piscivorous young-of-the-year bluefish iPomatomus saltatrix) in the Hudson River estuary. Fish. Bull. 95:665- 679. Buckel, J. A., D. O. Conover, N. D. Steinberg, and K. A. McKown. 1999a. Impact of age-0 bluefish iPomatomus saltatrix) pre- dation on age-0 fishes in the Hudson River estuary: evidence for density-dependent loss of juvenile striped bass. Can. J. Fish. Aquat. Sci. 56:275-287. Buckel, J. A., M. J. Fogarty, and D. O. Conover. 1999b. Foraging habits of bluefish (Pomatomua saltatrix) on the U.S. east coast continental shelf Fish. Bull. 99: 758-775. Buckel, J. A., N. D. Steinberg, and D. O. Conover. 1995. Effects of temperature, salinity, and fish size on growth and consumption ofjuvenile bluefish. J. Fish Biol 47:696-706. Clepper, H., ed. 1979. Predator-prey systems in fisheries management. Sport Fishing Institute. Washington, D.C., 504 p. Edwards, R. L., and R. E. Bowman. 1979. Food consumed by continental shelf fishes. In H. Clepper (ed.). Predator-prey systems in fisheries manage- ment, p. 387-406. Sport Fisheries Institute, Washington, D.C. Hartman, K. J., and S. B. Brandt. 1995a. Comparative energetics and the development of bioenergetics models for sympatric estuarine piscivores. Can. J. Fish. Aquat. Sci. 52:1647-1666. 1995b. Predatory demand and impact of striped bass, blue- fish, and weakfish in the Chesapeake Bay: apphcations of bio- energetics models. Can. J. Fish. Aquat. Sci. 52:1667-1687. 1995c. Trophic resource partitioning, diets, and growth of sympatric estuarine predators. Trans. Am. Fish. Soc. 124:520-537. Juanes, F., and D. O. Conover. 1994. Rapid growth, high feeding rates, and early pLscivory in young-of-the-year bluefish (Pomatomus saltatrix). Can. J. Fish. Aquat. Sci. 51:1752-1761. Kerfoot, W. C, and A. Sih. 1987. Predation:direct and indirect impacts on aquatic com- munities. Univ. Press of New England. Hanover. NH, 386 p. Lassiter, R. R. 1962. Life history aspects of the bluefish, Pomatomus saltatrix I Linnaeus), from the coast of North Carolina. M.S. thesis. North Carolina State College, Raleigh, NC, 103 p. Magnusson, K. G. 1995. An overview of the multispecies VPA-theory and applications. Rev. Fish Biol. Fisheries. 5:195-212. Munch, S. 1997. Recruitment dynamics of bluefish, Pomatomus saltatrix, on the continental shelf from Cape Fear to Cape Cod, 197.3-1995. M.S. thesis. State University of New York at Stony Brook, NY, 127 p. Oviatt, C. A. 1977. Menhaden, sport fish, and fishermen. University of Rhode Island Marine Technical Report 60, 24 p. Palomares, M. L., and D. Pauly. 1989. A multiple regression model for predicting the food consumption of marine fish populations. Aust. J. Mar. Freshwater Res. 40:259-273. Pauly, D. 1986. A simple method for estimating the food consump- tion offish populations from growth data and food conver- sion experiments. Fish. Bull. 84:827-839. Pauly, D., and M. L. Palomares. 1987. Shrimp consumption by fish in Kuwait waters: a meth- odology, preliminary results and their implications for man- agement and research. Kuwait Bull. Mar Sci. 9:101-125. Buckel et al.: A comparison of biomass harvested by Pomatomus saltatnx with that harvested by fisheries 785 Richards, S. W. 1976. Age, growth and food of bluefish from east-central Long Island Sound from July through November 1975. Trans. Am. Fish. Soc. 105:523-525. Sissenwine, M. P., E. B. Cohen, and M. D. Grosslcin. 1984. Structure of the Georges Bank ecosystem. Rapp. R- V. Reun. Cons. Int. Explor. Mer 183:243-254. Sissenwine, M. P., and N. Daan. 1991. An overview of multispecies models relevant to manage- ment of living resources. ICES Mar. Sci. Symp. 193:6-11. Stewart, D. J., and F. P. Binkowski. 1986. Dynamics of consumption and food conversion by Lake Michigan alewives: an energetics-modeling synthesis. Trans. Am. Fish. Soc. 115:643-661. Yanez-Arancibia, A., A. L. L. Dominguez, and D. Pauly. 1993. Coastal lagoons as fish habitats. /;; B. Kjerfve (ed.), Coastal lagoon processes, p. 339-351. Elsevier Science Publishers, Elsevier Oceanography Series 60, Amsterdam. 786 Abstract.— The abundance and orien- tation of trawl marks was quantified over an extensive portion (>2700 kni^) of the Eureka, CaHfornia, outer shelf and slope, an important commercial bottom trawling ground for such high- value species as rockfish, sole, and sablefish. Fishing logbook data indicate that the entire reporting area was trawled about one and a half times on an average annual basis and that some areas were trawled over three times annually. High-resolution sidescan- sonar images of the study area revealed deep gouges on the seafloor, caused by heavy steel trawl doors that act to weigh down and spread open the bot- tom trawls. These trawl marks are com- monly oriented parallel to bathymetric contours and many could be traced for several kilometers. Trawl marks showed a quadratic relationship in relation to water depth, with the greatest number of trawl marks observed at -400 m. There was a significant positive corre- lation between the number of trawl marks observed on the sidescan images and the number of annual trawl hours logged within reporting areas. This finding indicates that acoustic remote sensing is a promising independent approach to evaluate fishing effort on a scale consistent with commercial fish- ing activities. Bottom trawling gear is known to modify seafloor habitats by altering benthic habitat complexity and by removing or damaging infauna and sessile organisms. Identifying the ex- tent of trawling in these areas may help determine the effects of this type of fish- ing gear on the benthos and develop indices of habitat disturbance caused by fishing activities. Sidescan-sonar mapping of benthic trawl marks on the shelf and slope off Eureka, California Alan M. Friedlander Pacific Fisheries Environmental Laboratory Southwest Fisheries Science Center National Marine Fisheries Service, NOAA 1352 Lighthouse Avenue, Pacific Grove, California 93950 Present address: The Oceanic Institute Makapuu Point, 41-202 Kalanianaole Highway Waimanalo, Hawaii 96795 E-mail address afnedlandergteligentmail com George W. Boehlert Pacific Fisheries Environmental Laboratory Southwest Fisheries Science Center National Marine Fisheries Service, NOAA 1352 Lighthouse Avenue, Pacific Grove, California 93950 Michael E. Field United States Geological Survey, Coastal and Marine Geology 345 Middlefield Road, Menio Park, California 94025 Janet E. Mason Pacific Fisheries Environmental Laboratory Southwest Fisheries Science Center National Marine Fisheries Service, NOAA 1352 Lighthouse Avenue, Pacific Grove, California 93950 James V. Gardner Peter Dartnell United States Geological Survey, Coastal and Marine Geology 345 Middlefield Road, MenIo Park, California 94025 Manuscript accepted 7 April 1999. Fish. Bull. 97:786-801 ( 1999). Concerns about trawl impacts on benthic fish habitats date back to the earliest use of this gear in the 13"^ and 14"^ centuries (de Groot, 1984), but the extent and longevity of these impacts have been difficult to quantify. The effects of trawling depend on the size and type of bot- tom trawl, footrope gear, bridles, doors, scope of main wires, trawl- ing speed, duration, and repetition of trawling. Vulnerability of the sea floor to trawling impacts depends on the nature of the bottom type, benthic fauna, sedimentation rates, tidal velocity, and the degree of re- working of sediments caused by storms. Seabed disturbance can oc- cur over the entire distance between the doors but is most pronounced in the region scoured by the trawl doors (Messieh et al., 1991), which have been shown to plough to depths greater than 15 cm (Caddy, 1973: Churchill, 19891. In the short term, disturbance from bottom trawling can cause resuspension of sediments and make benthic infauna more avail- able to scavenging predators (Kai- ser and Spencer, 1994; Kaiser and Ramsay, 1997; Ramsay et al., 1997; Lindeboom and de Groot, 1998). Long-term shifts in abundance and Fnedlander et al : Sidescan-sonar mapping of benthic trawl marks off Eureka, California 787 diversity of benthic fauna have been noted in areas where trawling has been conducted for extended pe- riods of time (Reise, 1982; de Groot, 1984; Sainsbury, 1987; Hutchings, 1990; Colhe et al., 1997; Lindeboom and de Groot, 1998). Chronic, long-term disturbance can delay or prevent recovery of ecological commu- nities (Collie et al., 1997) and large-scale removal of macrobenthos can lead to permanent changes in the community (Jones, 1992; Collie, 1998; Rogers et al., 1998). The removal of benthic organisms and sedi- mentary structures (e.g. sand waves, depressions) by trawls can lead to modifications of the physical com- plexity of benthic habitats (Reise, 1982; Jones 1992; Auster et al., 1996; Collie, 1998; Dorsey and Peder- son, 1998), Schwinghamer et al. (1996, 1998) found that trawling homogenized sediments on the Grand Banks of Newfoundland to a depth of at least 4.5 cm, thus reducing fine-scale complexity of the substrate. Sediment mixing and frequent bottom disturbance from trawling activity may affect resuspension fluxes and produce changes in the successional organiza- tion of soft-sediment infaunal communities (Pilskaln et al., 1998). Reduction or removal of benthic struc- tural complexity can lead to reduced recruitment of benthic organisms and often to reduced production (Botsford et al., 1997). The degree of impact on benthic habitat is related to the timing, severity, and frequency of disturbance (Watling and Norse, 1998). The average annual area swept by trawls on Georges Bank from 1976 to 1991 was between 200% and 400% of the total area (Auster et al., 1996). Trawling does not occur evenly over this area but is concentrated in locations that produce better catches and fewer obstructions to towing (Dorsey and Pederson, 1998; Auster and Langton, 1999). As a result, some areas are subjected to trawl- ing at a much higher rate than these reported aver- ages and other areas may not be trawled at all. The effects of dredging on tidal flats in the Wadden Sea persisted for more than 15 years (van der Veer et al., 1985); by contrast, experimental trawl marks in a relatively dynamic intertidal zone in the Bay of Fundy persisted only 2-7 months (Brylinsky et al., 1994). The effects of mobile fishing gear on bio- diversity are most severe where natural disturbance is least prevalent, such as at outer continental shelf and slope habitats (Watling and Norse, 1998). An experimental program to examine the impacts of mobile fishing gear on the benthic ecosystems in At- lantic Canada clearly indicated that trawling changed the physical habitat structure on sandy bot- tom at 120-146 m over a three-year period (Gordon et al., 1998). Biomass of epibenthic organisms in the trawl catch decreased with repeated trawling and the total biomass, as sampled by epibenthic sled, was lower in trawled areas compared with adjacent habi- tats. Fishing effort is spatially nonuniform, and thus techniques are needed that will allow analysis of trawling activity, and thus trawling impacts, on spa- tial scales appropriate to the large scales on which fisheries operate. In this study, we examine sidescan- sonar records as one approach to this analysis. The continental shelf off Eureka (water depths less than -120 m) is relatively flat (<0.5 degree) and topo- graphically very smooth due to large inputs of fine terrigenous sediments from the Eel and Mad Rivers and reworking by storms (Fig. 1; Goff et al., 1999). Bottom photographs and samples from the shelf gen- erally show smooth, mud-covered bottom with small- scale bottom roughness resulting from bioturbation and occasional wave ripples (Wiberget al., 1996). The dominant morphological feature of the shelf is the Eel River delta, a sediment bulge in water depths of 20 to 60 m adjacent to the mouth of the Eel River and elongated to the north. Slope deposits are domi- nated by fine-grain sediments that show a transi- tion from sandy silt in the south near the Eel Can- yon to clayey silt in the northern portion of the study area (Syvitski et al., 1996). Seafloor depressions 5 to 20 m in diameter and sev- eral meters deep, termed "pock marks," are common on both the outer shelf and slope throughout the re- gion. These depressions are inferred to form by the re- moval of particles during expulsion of gas and fluid fi"om subsurface deposits (Yun et al., 1999). The excavation is thought to occur both at slow rates, inducing par- ticle-by-particle removal, and at rapid rates during sudden release of subsurface gas by earthquake shak- ing or other mechanisms (Field and Jennings, 1987). In addition to natural processes that modify the seafloor, such as bottom currents, landslides, and fluid-expulsion features, the sidescan data and bot- tom photographs also show evidence (trawl marks) of fishing activities that modify these benthic habi- tats. The northern California outer continental shelf and slope off Eureka are important commercial fish- ing grounds. The trawl fleet that fishes the area off Eureka presently consists of 36 vessels from Eureka and 29 from Crescent City.^ These vessels range in size from 18 to 24 m (60 to 80 ft), the larger vessels having entered the fleet in recent years in response to a shift of fishing effort into deeper water. Trawl designs vary but currently the most common trawl configuration is a Nor'eastern design with 114-mm (4 1/2 in) diamond stretched-mesh polyethylene net- ting equipped with roller gear. Trawls are currently Quirollo, L. F. 1998. California Department of Fish and Game, 619 Second Street. Eureka, CA 95501. Personal commun. 788 Fishery Bulletin 97(4), 1999 X/1^ Figure 1 Perspective, shaded-relief view of the study area based on high-resolution multibeam bathymetry. Vertical exaggeration is 20-. The continental shelf (< 120 m water depth) is bathymetrically smooth owing to large input of fine terrigenous sediments from the Eel and Mad Rivers and to reworking by storms. The southern portion of the shelf is dominated by the Humboldt slide. The northern portion of the slope is -60 m deeper and ~7 km farther from shore than in the Humboldt slide area. rigged with a 33.5-m { 110-ft) net opening with trawl doors hung -76-91 m (250-300 ft) apart. The trawl fishery on the shelf primarily occurs over soft-sediment habitat and mainly targets flatfishes. The important species in this assemblage include English sole (Pleuronectes vetulus), Petrale sole (Eopsetta jordani), and sanddabs iCitharichthys spp. ). The slope fishery occurs over rougher and more heterogeneous habitats that begin at the shelf break (-120 m) and continue to depths >900 m. The catch in this assemblage is dominated by Dover sole (Mi- crostomus pacificus), sablefish (Anoplopoma fimbria ), thornyheads (Sebastolobus spp.), and several rock- fish species (Sebastes spp.). A midwater trawl fish- ery also exists in waters above the slope for Pacific whiting (Merliiccius productiis) and widow rockfish {Sebastes entomelas). The assemblages in this region are similar to those found along other portions of the U.S. west coast by the National Marine Fisheries Service triennial trawl surveys (Jay, 1996). Groundfish are among the most diverse and eco- nomically valuable fishery resources along the west coast of the United States (NOAA, 1996). Diversity, quality, and extent of habitat are among the most important environmental determinants of distribu- tion, abundance, and species diversity for ground- fish (Carlson and Straty, 1981; Matthews and Richards, 1991). The broad spatial extent of these fisheries generally precludes careful examination of the nature of the exploited habitats, the relationship among species and habitats, and the degree to which fishing activities have affected these habitats. Con- servation of fisheries habitat is an important con- sideration for sustaining fisheries production. The reauthorization of the Magnuson-Stevens Fisheries Conservation and Management Act requires incor- porating the concept of "essential fish habitat" in Fishery Management Plans (Schmitten, 1996). Be- cause fishing, particularly with bottom trawls, can alter essential fish habitat, it is important to quan- tify fishing activity and its effects on the associated habitat. The goals of this study are to evaluate the extent of marks from bottom trawls off Eureka, Cali- fornia, at a scale consistent with commercial fishing Fnedlander et al.: Sidescan-sonar mapping of benthic trawl marks off Eureka, California 789 activities and to compare these results with fishing- effort data from commercial fishery logbook data. Materials and methods Geological setting The study area is the upper continental margin of northern California, north of Cape Mendocino. The shelf and slope in this region are within the Eel River Basin, a forearc basin filled with sediment derived from high sediment loads during the late Neogene (Clarke, 1992). The entire basin is highly faulted and folded from multiple episodes of deformation. Shear- ing and compression of the margin related to under- thrusting of the Gorda plate beneath the North American plate has resulted in thrust faulting, up- lift, and an overall youthful geologic setting, and this has produced varied relief and structure of the sea- floor (Carver, 1987; Clarke and Field, 1989). Studies of seafioor stiaicture and sediment history of this area (Field et al., 1999; Gardner et al., 1999; Goff et al., 1999) have identified many features related to submarine landslides and to downslope sediment trans- poi't through channels and gulleys. Sidescan-sonar images of the seafloor along the shelf and slope, as well as multibeam maps, high-resolution seismic I'efiection data, and instrumented tripods from complementaiy investigations, all show abundant evidence of modern processes at work on the seafloor (Field and Jennings, 1987; Alexander and Simoneau, 1999; Field et al., 1999; Gardner etal., 1999; Goff etal., 1999;Yunet al., 1999). There are three main types of relief on the sea- floor in this region. Along the slope, in the center of the study area, a large growth fold (anticline) has uplifted through the modern seafloor, exhibiting 100 m of steep relief. South of the anticline, the slope between 120 and 600 m depths is dominated by the Humboldt slide, an area where slope deposits have undergone deformation and limited downslope move- ment (Field and Barber, 1993; Gardner et al., 1999). North of the anticline, the slope is crossed by a series of shallow gullies that are 100 to 200 m wide and 1 to 2 m deep (Field et al., 1999). Gullies are spaced about 200 to 300 m apart. Gullies also occur south of the anticline where they are generally fewer in num- ber and somewhat deeper (-20 m). The gidlies are thought to represent sediment pathways during ear- lier periods when sea level was at a lower position. Sidescan survey - Through a collaborative effort of the U.S. Geological Survey (USGS) and the National Marine Fisheries Service (NMFS), we examined sidescan-sonar data off Eureka, California, collected by the USGS during the STRATA FORMation on Margins (STRATAFORM) progi-am sponsored by the Office of Naval Research (Nittrouer and Ki-avitz, 1996; Nittrouer, 1999). A grid of 1600 line kilometers of high-resolution, digital, deep-towed (95 kHz) sidescan-sonar data was ob- tained during 1995 and 1996 (Fig. 2). The sidescan- sonar images were obtained by using the USGS Datasonics SIS- 1000 chirp sonar system and an ISIS data-acquisition system. The sidescan sonar operates as a swept FM signal with a frequency band from 90 to 110 kHz (port channel sweeps from low to high, starboard channel sweeps high to low). Usual swath width during the cruises was 0.5 s (750-m swath). Sidescan data were recorded on magneto-optical disks and downloaded to a DEC ALPHA workstation. The digital sidescan-sonar data were all processed with USGS MIPS image-processing software (Chavez, 1984; 1986). The raw data were geometri- cally corrected (slant range) by using the position, heading, speed, and nadir depth of the ship. Each of the 8-bit image pixels was geometrically located by assuming a flat seafloor which introduces a certain amount of uncertainty into the actual location of any given pixel. Our estimated location accuracy was about 100 m. After the geometric corrections were made, anamorphic corrections were applied to ac- count for the aspect ratio between along- and across- track directions. Radiometric corrections had to be applied to the 8-bit digital number (DN) that repre- sents backscatter. Each sidescan line was subdivided into segments and, for each segment, the average DN for each binned range was normalized to the mean of all the binned ranges within that segment. Each segment was several hours long, and this cor- rection produced tone-matched adjacent segments. Trawl marks Ti'awl marks appear on the sidescan sonographs as long, narrow, linear depressions (Fig. 3 ). Trawl marks were traced on mylar overlaid on the sidescan records. Each sidescan track line was divided into 10-min time intervals. Vessel speeds during the sur- veys varied between 3 and 4 knots, and therefore each 10-minute interval covered from 0.9 to 1.2 km along the ship's course. The sidescan record covered a 750-m swath so that 10-min segments represent from 675,000 m- to 900,000 m- of area. To quantify the number of trawl marks per time interval, lines were drawn perpendicular to the course of the side- scan track line, and trawl marks that bisected these lines were counted. Mean trawl mark length was 0.77 km (SD=0.26) for 20 randomly selected 10-min 790 Fishery Bulletin 97(4), 1999 1 24''20' 124°10' 124° 41°10' 40°50' 40°40' 40°40' 40°30' 40°30' 124°30' 124°20' 5 124° 10' 10 nautical miles 10 15 20 l^ilometers Figure 2 Track lines for high-resolution sidescan-sonar surveys. The reporting blocks from California Department of Fish and Game that are used for logbook data entry are superim- posed on the sur\'ey area. Blocks are 10 minutes of longi- tude and 10 minutes of latitude on a side with block num- ber in the center. intervals from 10 different sidescan track lines (Fig. 4). From these results, it was determined that two random perpendicular lines were needed to prop- erly sample each 10-min interval and minimize double-counting of trawl marks. Using a sample of 50 randomly selected 10-min time intervals from 10 sidescan track lines, we counted the actual number of trawl marks and compared this number with the mean number counted with the perpendicular line intercept method described above. The mean propor- tion of trawl marks counted with this sampling method to actual number of marks was 0.84 (SD=0.11). Trawl mark density in kni'^ was calculated from each time interval and incorpo- rated into a Geographical Information System (GIS) for analysis and contouring. The angle of each trawl mark was measured in relation to the two perpendicular lines used to quantify trawl marks in all 10-min inter- vals. These angles were adjusted for the course of the sidescan track and computed in rela- tion to magnetic north. Mean angles for each time interval were then calculated and plot- ted on a polar plot in an r = /! 6) format, where r is the number of trawl marks and theta (6) is the angle relative to 0°. The grand mean angle and 95% CI were computed with the mean values from each time interval (Zar, 1984). Fishing effort Trawlers provide information on catch and effort of individual trawl hauls to the Califor- nia Department of Fish and Game (CDF&G). Hauls are spatially recorded in blocks that are 10 minutes of latitude by 10 minutes of longi- tude (Fig. 2). Blocks within the Eureka STRATAFORM survey area were extracted from the NMFS Tiburon Laboratory ground- fish database and used in these analyses. Trawling is prohibited by state law within 5.6 km (3 nm) of the coast. All reporting blocks adjacent to the coast were truncated for all estimates and comparisons. Pacific whiting form extensive midwater ag- gregations during the day, and the fishery is conducted almost exclusively with midwater trawls (Dorn and Saunders, 1997). Because fishermen did not report whether their trawl hauls were bottom trawls or midwater trawls, all trawl hauls with catches of Pacific whiting greater than 454 kg ( -1000 lb) were eliminated from these analyses. Mean annual fishing ef- fort (trawl hours) was calculated for each re- porting block within the Eureka area for the years 1990 to 1994. Depths from the logbook data were used to obtain mean fishing depths for each reporting block for comparisons with fishing effort. The reporting block grid was incorporated into a GIS and the mean density of trawl marks was calcu- lated for each block (Table 1). Water depths from a high-resolution multibeam mapping sui"vey (Goff et al., 1999) were binned to obtain mean bottom depth in each block for comparison with mean number of trawl marks. Effort, expressed as mean annual trawl Fnedlander et al.; Sidescan-sonar mapping of benthic trawl marks off Eureka, California 791 hours per block, was compared to the mean number of trawl marks in each reporting block with a Spearman rank order correlation. Two reporting blocks (129 and 218) were excluded from this analysis owing to a lack of adequate sidescan coverage. To estimate the total area swept by bottom trawls on an annual basis. the total annual number of fishing hours per reporting block was multiplied by a typical vessel speed (5.5 km/h) and a typical door to door width of 85 m. Detrended correspondence analysis (DCA) was used to identify clusters of similar report- ing blocks in ordination space on the basis of fish assemblage structure from CDF&G log- book data (Ludwig and Reynolds, 1988). A matrix of reporting blocks by catch was cre- ated for use in this analysis. Habitat types were defined a priori by depth and physical characteristics of the habitat from previous geological surveys and then overlaid on the station clusters created by DCA. A Kruskal- Wallis rank sum test (Hollander and Wolfe, 1973) was used to compare the density of trawl marks and total fishing hours per reporting block in each habitat type discerned by DCA. Pairwise comparisons of trawl marks and fish- ing effort between habitats was conducted by using Dunn's multiple comparison procedure at a = 0.05 (Hollander and Wolfe, 1973). Results We measured the range, density, and orienta- tion of marks on the seafloor caused by trawl- ing activity as resolved by the sidescan-sonar records. Of the 1246 10-min intervals exam- ined, -10% were washed out and could not be analyzed. Densities of trawl marks/km- per 10- min interval ranged from 0.00 to 98.55 and mean density per block ranged from 0.94 to 38.3 with a grand mean of 20.0 marks/km- (Table 1). Trawl marks were most abundant on the slope, particularly the northern portion and least abundant on the con- tinental shelf (Fig. 5). Trawl marks commonly were oriented parallel to isobaths and some could be traced for several km (Fig. 6). The overall mean trawl direction was 352.6° (95^7^ CI -i-/-5°) (Fig. 7) which agrees with the general ori- entation of isobaths in this area. Because trawl marks were generally orientated in the north— south direc- tion along isobaths, sidescan tracks orientated in the east-west direction (>45° and <135°) might not dis- tinguish as many trawl marks in relation to those Figure 3 Sidescan-sonar image (white is shadow; dark is reflection ) of trawl marks on the seafloor at a water depth of ~ 150 m. Sidescan swath width is 750 m. Tick marks are .5-min time intervals (-0.55 km). Arrows point to examples of trawl marks. Shallow circular depres- sions on the sonograph are gas pockmarks. Ship's course for this sonogram is 345". orientated in the north-south direction. Using 25 randomly selected time intervals from five east-west sidescan track lines and 25 from five north-south track lines, we found that the proportion of trawl marks sampled with the perpendicular line intercept method was not significantly different from the ac- tual number of trawl marks for the two track line orientations (Mann-Whitney rank sum test=759.5, P=0.775). We compared the orientation of trawl mark angles between east-west and north-south orientations by selecting seven locations where east-west and north- south sidescan-sonar track lines intersected during the 1996 survey. The intersecting time interval, along 792 Fishery Bulletin 97(4), 1999 with one time interval to either side were pooled for each orientation (21 time intervals for each orienta- tion) and compared by using a Watson-Williams two- sample test for mean angles (Zar, 1984). There was a significant difference (F=44.7, P<0.001 ) in the mean angle of trawl marks detected along north-south Table 1 "Summary statistics for density estimates of trawl marks in each California Department of Fish and Game (CDF&Gl reporting block. Block numbers with asterisks are excluded from comparative analysis with fishing effort owing to small sample size and lack of adequate block coverage. CDF&G block Mean (no/km^) SD (no/km^) Max no/km^) Min (no/km^ ) n 127 2.5.22 16.77 52.92 0.00 34 128 3.5.07 18.56 98.55 2.09 132 129* 31.19 7.94 45.08 21.59 13 133 11.80 13.20 49.13 0.00 81 134 38.26 11.82 63.49 0.00 144 135 22.02 10.69 52.11 1.39 34 202 4.13 6.44 26.67 0.00 169 203 18.01 12.65 43.87 0.00 159 204 22.82 15.71 64.09 0.00 115 210 0.94 2.25 14.03 0.00 69 211 10.99 16.89 76.75 0.00 102 212 28.98 13.56 66.42 8.25 39 217 27.39 27.23 75.55 0.00 25 218* 28.68 15.58 39.69 17.66 2 Mean 19.99 18.34 98.55 0.00 1= 1118 E S Z 0 10 A/ =405 Mean = 0.77 km SD = 0.26 n rx Length (km) Figure 4 Frequency distribution of trawl-mark length per 10-min interval. Ten minute time inters'als ranged from 0.77 to 1.2 km in length (mean=0.91 ). Data are from 20 randomly selected 10-min intervals from 10 different sidescan tracks. orientated track lines (mean=357.91 compared with east-west orientated track lines (mean=294.5) at these intersection points. The mean density of trawl marks in each CDF&G reporting block fitted a quadratic relationship with water depth (y=-0. 0002x2 + 0.1374x -i- 2.9353, 7?2=o.501, P=0.64). The lowest density of trawl marks was observed in the shallowest water with maximum number of trawl marks observed at ~400m depth (Fig. 8). Mean annual trawl hours ( 1990-94) also fit- ted a quadratic model with water depth for each re- porting block {y=-0.Q082x'^ + 8.7015x - 960.57, i?2=o.723, P<0.001) (Fig. 8). There was a significant positive correlation between the density of trawl marks and the mean annual number of trawl hours per reporting block (r,=0.573, P=0.048). The fishing blocks off Eureka can be separated into three depth zones on the basis of fish assemblages analyzed by DCA (Fig. 9). The upper and mid slope blocks are relatively close in DCA ordination space, whereas the shelf blocks showed a distinct assem- blage structure. The shelf blocks nearest the coast- line have effective fishing depths of <102 m (Table 2). From logbook data, the dominant species at these depths are flatfishes (e.g. sanddabs, Petrale sole, and English sole) and rockfishes. Dominant species from trawling on the upper slope ( 175 to 437 m) primarily consisted of Dover sole and rockfishes. Thornyheads, Dover sole, and sablefish are the dominant species on the mid slope (effective fishing depth from -595 to 837 m). The mean density of trawl marks was significantly different (Kruskal-Wallis //=287.4, P<0.001) among the three depth zones recognized by DCA (Fig. lOA). Upper and mid slope depths were not significantly different from one another (P>0.05) but both had signifi- cantly greater numbers of observable trawl marks than the shelf depth zone (P<0.05). Mean annual fishing effort per block also showed significant differences among depths (Kruskal-Wallis //=9.5, P==0.009, Fig. lOB). Fishing effort was lowest in the shelf habitat and highest in the upper slope, although the upper slope and deep slope depth zones were not significantly differ- ent fi-om one another (P>0.05). The cumulative annual area swept by commercial bottom trawls, as calculated from logbook data, was compared to the total area in each reporting block (Fig. 11). The entire study area was trawled -1.5 times on an average annual basis. The area trawled exceeded the total block area in the majority of the blocks. This Fnedlander et al : Sidescan-sonar mapping of benthic trawl marks off Eureka, California 793 was particularly true in the deeper blocks where the ratio of area trawled to block areas ranged from 127% to 339%. The mean trawl depth and mean duration of hauls for bottom trawls off Eureka has in- creased steadily since the early 1980s (Fig. 12A). Along with this change in fishing depth has come a shift in the composition of species landed. Thornyheads are deeper-water species and their catch contribution has increased from -4% of the total catch in the late 1970s to -14% in the 1990s. Conversely, the English sole is a shallow-water flatfish whose landings have declined from nearly 77c of the total catch in the late 1970s to 2-3% in the 1990s. Discussion 124°40' 4r20' 124°30' 124.°20' 124°10' Utility of sidescan sonar 41''10' 40°50' Acoustics are useful in fisheries for stock as- sessment (Karp, 1990), for remotely sensing characteristics of bottom habitat (Able et al., 1987; Greenstreet et al., 1997; Mayer et al., 1997; Yoklavich et al., 1997; Auster et al., 1998a, 1998b), and for assessing impacts to fisheries habitat (Auster et al., 1996; Collie et al., 1997). Our results demonstrate that the use of acoustic remote sensing also presents a promising independent approach for evaluat- ing fishing effort on a spatial scale consistent with commercial fishing activities. Sidescan- sonar data are useful for identifying the loca- tion of trawling activity, as well as for evalu- ating the amount of disturbance to the sea- bed, and may be used in developing indices of habitat disturbance caused by fishing activi- ties. Despite minimal sidescan coverage in some reporting blocks and biases in the log- book data associated with misreporting, there was a significant positive correlation between the density of trawl marks per reporting block from sidescan-sonar data and the mean annual number of trawl hours. A quadratic relationship with depth was observed for the number of trawl marks and fish- ing effort per reporting block with maximum values at -400 m and -500 m. respectively. This relation- ship, together with the density contour plot of trawl marks and the vector plot of trawl orientation, shows that fishing effort was concentrated on the upper to mid slope and along bathymetric contours. The entire study area off Eureka is trawled -1.5 times its total area on an annual basis. Because of narrow bands of rocky outcrops and anticlines. 124° 41°20' 41°10' 41° 40°50' 40°40' 40°40' 40°30' 124°40' 124°30' 124°20' 124°10' 124° 40°30' Trawl marks/sq. km 0-10 0 5 10 nautical miles 0 5 10 15 20 kilometers Figure 5 Contour plot of trawl mark densities (number/km-) de- rived from 10-minute intervals along sidescan survey track lines. Actual locations of data points are in Fig. 6. trawlable fishing area within the mid slope report- ing blocks is less than the total area of the blocks. The area swept by trawls is therefore an under-esti- mate of the effective fishing area that is impacted by trawl gear in these blocks. Gislason ( 1994) estimated that trawling swept the entire area of the North Sea annually, with some fishing grounds in the southern North Sea likely being swept more than three times a year by beam trawls. This type of chronic impact on the seafloor can have a long-term effect on the local benthic epifauna (Rogers et al., 1998). For ex- ample, there is abundant evidence of rich ecological communities associated with cold-seeps and exten- 794 Fishery Bulletin 97(4). 1999 sive bioturbation off Eureka (Syvitski et al., 1996). Released gases and fluids from cold seeps are rich in methane and sulfide and support unique productive cold-seep biological communities (Syvitski et al., 1996). The magnitude of trawling activity in this area has the potential to modify these communities by direct physical damage or by smothering from resus- pended sedimention. There has been increased interest in recent years on the impacts of fisheries on the marine environ- ment (Dayton et al., 1995; Boehlert, 1996; Lindeboom 124°30' 124°20' 124° 10' 124° 40°50' 40°40 40°40' 40»30' 40°30' 124° 10 IS 20 kilometers Figure 6 Vector plot .showing orientation and magnitudi' of trawl marks along sidescan track lines. Length of arrow is pro- portional to the density of trawl marks in each 10-min interval, and direction is the mean angle of all trawl marks in each time interval. Note that angles are plotted in re- lation to north so that trawls towed north or .south resulted in the same orientation. and de Groot, 1998; Watling and Norse, 1998), and an important component is the relationship between exploited fish species and their associated habitats (Schmitten, 1996). With a greater understanding of these associations, concerns have been raised about the adverse impacts that mobile fishing gear may have on the entire ecosystem. The severity and du- ration of disturbance are both factors that affect eco- logical communities (Gislason, 1994; Connell et al., 1997; Tuck et al, 1998; Auster and Langton, 1999). Mobile fishing gear, such as trawls, is the most wide- spread form of direct disturbance to marine systems below depths affected by storms (Watling and Norse, 1998). Natural distur- bances to the benthic community off Eureka probably occur from high-energy wave events, floods, and even tectonic activity. The first two processes likely have a greater impact in the shallower water on the shelf, and the latter pro- cess is likely too infrequent to play a major role in impacting the benthos. Thus, in the habitat covered by this study, trawls represent the greatest form of direct disturbance to the benthic habitat — a finding that is consistent with the conclusions of Watling and Norse (1998) for depths below the influence of storms. The extent and location of fishing activity, noted in this study from logbook data and sidescan- sonar records, can potentially have a large im- pact on the local ecological communities, par- ticularly on the slope where the frequency and magnitude of natural disturbance is less than that on the shelf Examination of trawl marks was not the pri- mary goal of the sidescan-sonar survey con- ducted off Eureka (Syvitski et al., 1996). Con- sequently, there are aspects of the survey method that could have been conducted differ- ently to improve the quality of these data. Be- cause fishing activities in this area are gener- ally conducted along the north-south bathymet- ric contours, sidescan-sonar tracks oriented east-west tended to resolve slightly fewer trawl marks. Any future study of trawl marks using sidescan sonar should concentrate tracks to ori- entations subparalleling trawling directions in order to maximize the resolution of the survey. There are biases associated with this method, particularly in a dynamic physical location like the northern California continental margin. Estimates of the actual number of trawl marks are minimal because of variability in resolution of the sidescan records, obliteration of older trawl marks by more recent trawling, burial by sedimentation or reworking of the seafloor by Friedlander et al : Sidescan-sonar mapping of benthic trawl marks off Eureka, California 795 270 90 Figure 7 Polar plot of mean angle and density of trawl marks for each time inter- val. Polar plot in an r = /IS) format where r is the density of trawl marks and theta (6) is the angle relative to 0 = . The grand mean trawl direction was 352.6" OSVf CI +/-5'). large waves, particularly at shallower depths. Trawl marks may persist for longer duration in deeper water where sedimentation rates are lower and storm waves have a lessened or neghgible effect. Shelf sedi- ment presently accumulates 2 to 10 times faster on the shelf than on the slope ( 1-3 mm/yr on the slope; 6-10 mm/yr on the shelf based on -lopB activities: Alexander and Simoneau, 1999; Sommerfield and Nittrouer, 1999). At present rates, and assuming trawl track depths of -15 cm, it would take 50-75 years to fill in a typical trawl mark, but evidence shows that small natural gullies on the slope are being draped and are infilled at present and thus maintain their shape and relief even though they are in the process of being buried (Field et al., 1999). The sedimentation rates on the shelf are higher than on the slope (the flood deposits from the 1964 Eel River flood now lies -45 cm below the seafloor) (Sommerfield and Nittrouer, 1999), and annual storm waves resuspend sediments at depths up to 100 m. Resuspension and redeposition of silt particles may lead to draping of trawl gouges and thus preserve their shape and relief on the seafloor. In the pres- ence of bottom currents, which usually are present from tides, storm waves, or boundary flows, the re- suspended particles may fill in the trawl gouges and thus slowly erase them over time. Since fishing ef- fort in the study area was shown to be concentrated in water deeper than 100 m, these effects are possi- bly compounded by the longevity of trawl marks in this habitat. Research is needed to address the lon- gevity of trawl marks in various environments as well as to determine the impacts on (and recovery of) di- versity and production in benthic communities. 3000 1C00 - 1 B i fl2 = 0.723 P=<0.001 } y i i ^~~^\ - A 200 400 600 Mean depth (m) 800 Figure 8 (A) The nonlinear relationship between mean den- sity of trawl marks per 10-nim interval by CDF&G re- porting block depth ( m ). Mean trawl mark block depth is calculated from the mean depth of 10-min intervals along the sidescan track in each reporting block. (B) Mean annual fishing hours per reporting block and block depth (m). Mean fishing hour block depth is calculated from the depths reported from logbooks in each report- ing block. Error bars are standard error of the mean. 796 Fishery Bulletin 97(4), 1999 400 1 300- 200 = 100 w < 0 -100 -200 -300 Upper slope Middle slope Shelf -200 I -100 100 Axis I 200 I 300 —I 400 Figure 9 Results of a detrended correspondence analysis of mean annual species catch compo- sition ( 1990-94) in CDF&G reporting blocks off Eureka. Three clusters are recog- nized as unique depth zones. Numbers are CDF&G reporting block numbers. Table 2 Characteristics of California Department of Fish and Game (CDF&G) reporting blocks by habitat zone. Values for fishing depth are mean depth per trawl in each reporting block from logbook data (1990-94). Block depth and area are calculated from Geo- graphical Information System (GISl data and exclude areas within the 5.6-kni (3-nmil no trawl zone. Values for dominant species are percentages of mean total annual catch (1990-94). Habitat zone Block Fishing depth (m) Block depth (m) Block area (km-) Fishing effort Dominant species caught Shelf 127 68 83 251 191 Sanddab 30.31 English sole 27.59 Petrale sole 14.82 133 102 83 207 16 Sanddab 33.31 English sole 14.35 Widow rockfis 1 9.87 202 67 69 202 169 Sanddab 48.46 English sole 12.80 Petrale sole 11.88 210 93 52 55 78 Rockfish 29.46 Sanddab 25.85 Petrale sole 14.25 217 101 96 147 170 Widow rockfi sh 52.36 Rockfish 12.64 Sanddab 11.29 Zone mean 86 77 I = 862 1 = 623 Sanddab 29.84 Rockfish 13.80 English sole 13.15 Upper slope 128 359 347 259 691 Dover sole 40.10 Rockfish 22.78 Sablefish 13.92 134 437 425 259 1156 Dover sole 60.50 Sablefish 16.48 Thornyhead 7.76 203 265 260 260 1425 Dover sole 34.07 Rockfish 21.41 Petrale sole 10.29 211 175 118 256 599 Dover sole 22.68 Rockfish 22.52 Sanddab 13.49 Zone mean 309 2881 = 1034 y = 3871 Dover sole 39.34 Rockfish 10.18 Sablefish 10.95 Mid slope 135 837 873 259 707 Thornyhead 45.75 Dover sole 33.11 Sablefish 16.21 204 539 583 260 1054 Dover sole 57.06 Sablefish 15.45 Thornyhead 12.13 212 525 570 260 1900 Dover sole 51.92 Thornyhead 16.80 Sablefish 15.63 Zone mean 634 675 I = 779 1 = 3661 Dover sole 47.36 Thornyhead 24.89 Sablefish 15.76 Friedlander et al : Sidescan-sonar mapping of benthic trawl marks off Eureka, California 797 The general temporal trend in fisheries develop- ment begins with localized fisheries that then expand geographically and to less accessible habitats, such as the deep sea (Deimling and Liss, 1994). In the present study, mean trawl depths have increased over the years as a result of changes in management strat- egies, changes in market demands, and overfishing of nearshore resources. Shortspine thornyheads {Sebastolobus alascaniis) and longspine thornyheads (S. altivelis) are slope rockfish species whose land- ings have increased along the West Coast in recent years (lanelli et al., 1994) whereas English sole has decreased. An export market for thornyheads devel- oped during the 1980s because a similar and highly valued species (S. macrochir) was depleted off Ja- pan (Rogers et al., 1997). As the Japanese market developed, the trawl fishery moved into deeper wa- 30 . A B B to 25 - T K-W = 287 4 T trawl mari o P< 0,0001 O 15 ■ 5- o 10 T3 A 03 c -r 1 ^ 0 - W=353 N = 562 N= 188 5000 ^ % 4000 K-W = 9 5 T 1 % P= 0.009 1 fishin a i 2000 Mean a A T N= 4 N = 3 1 N=5 Shelf Upper slope Middle slope Depth zone Figure 10 (A) Mean density of trawl marks per 10-min inten-al for the three habitat zones recognized by DCA. A^ = the num- ber of time intervals in each habitat zone. iBl Mean an- nual fishing hours per CDF&G reporting block for the three habitat zones recognized by DCA. TV = the number of re- porting blocks in each habitat zone. Error bars are stan- dard error of the mean. Statistical results from Kruskal- Wallis rank sum test. Dunn's multiple comparison proce- dure, a - 0.0.5. Habitats with the same letter designation are not significantly different from one another. ter following a typical pattern (Deimling and Liss, 1994) where longspine thornyheads and larger shortspine thornyheads are most common. Several of the most important species in the trawl catch (e.g. Dover sole and sablefish) segregate at depth by sex, size, and maturity stage. These biological character- istics as well as size and harvest restrictions makes interpretation of changes in the catch of these spe- cies difficult to interpret. Implications for management The groundfish trawl fishery has experienced declin- ing catches and increasing regulation in recent years. Current management strategies are imposed prima- rily on individual species or among small groups of congeneric species. These strategies do not address the impact to essential fish habitat from fishing ac- tivities and therefore may not be appropriate for the long-term sustainability of these resources. Marine reserves or harvest refugia are an effective manage- ment strategy that can help protect and maintain the complexity and quality offish habitat as well as mitigate the direct effects of fishing (Bohnsack, 1996; Bohnsack and Ault, 1996; Auster and Shackell, 1997; Yoklavich, 1998). A number of studies have noted reduced diversity and abundance of emergent epi- fauna in trawled areas compared with adjacent ar- eas closed to fishing (Bradstock and Gordon, 1983; Auster et al., 1996). The demersal trawl fishery off northwest Australia was thought to be responsible for a significant loss of emergent epifauna and decreased fish productivity (Sainsbury, 1987, 1988), both of which recovered in an area closed to fishing. The establishment of harvest refugia may be the most effective method of reducing the impact of trawls to benthic habitat and subsequently may help to improve the sustainability of the associated fish assemblages. Detailed analysis of sidescan-sonar data has al- lowed us to map and quantify trawl marks in rela- tion to fishery and habitat distributions. The combi- nation of geophysical data with fisheries dependent and independent data in a GIS provides the ability to examine the impact of trawling at a scale appro- priate for fisheries management. Evaluation of benthic habitat disturbance caused by fishing activi- ties is important for meeting the mandate of the Sustainable Fisheries Act. Developing indices of benthic habitat disturbance can be valuable in comparing the impact of fishing activities among dif- ferent areas and in establishing baselines for evalu- ating future management strategies such as closed areas. 798 Fishery Bulletin 97(4), 1999 124°30' 124°20' 124°10' 124° 41''10' 40°50' 40°40' 40°30' 41°10' 40°50' 40°40' 40°30' 124°30' 124°20' 124=10' S 10 nautical miles r ^ A ^ 400 y~\. - Depth >>-*< /^*^ ^ Hours ^ ^^"'^ £ / y-i Q. J i^ •D 300 j /A Mean fishing f Mean traw o o ^7 ' lours per haul 1 1 0 16 B /^x / \ B °'2- English sole / \ /" ' "1 Thornyheads / V^ Proportion of t o o g g 1 / 1976 1980 1984 1988 1992 1996 Year Figure 12 (A) Mean annual trawl depth and mean fishing hours | per trawl haul off Eureka from 1978 to 1992. Error bars are standard error of the mean. (B) Proportion of total groundfish trawl catch of thornyheads and English sole from 1978 to 1994. 10 IS 20 kilometers Figure 11 Total area trawled within each reporting block as a percentage of the total block area. Area trawled was calculated from the product of mean annual fishing effort 1 1990-94 1 from logbook data, typical vessel speed (5.5 km/h), and a typical trawl door to trawl door width of 85 m. "No trawl" areas within 5.6 km (3 nmii of shore are not included in the total area calculations. innumerable questions on the current status of the fishery. Don Pearson of the NMFS Tiburon Labora- tory was extremely helpful in providing California Department of Fish and Game commercial trawl log- book data for the Eureka region. Tone Nichols, formerly of the Pacific Fisheries Environmental Laboratory, as- sisted with earlier versions of several of the figures. This paper benefited greatly from comments provided by Paul Dayton. Ed DeMartini, Mary Yoklavich, Mark Zimmerman, and three anonymous reviewers. Acknowledgements This work was performed while the senior author held a National Research Council (NOAA/NMFS/ Pacific Fisheries Environmental Laboratory) Re- search Associateship. The Office of Naval Research (ONR Grant N00014-97-F0022 to M. E. Field) spon- sored the sidescan-sonar surveys conducted off Eu- reka during the STRATAFORM program. Larry Quirollo of the California Department of Fish and Game provided a great deal of insight into the his- tory of the Eureka trawl fishery as well as answered Literature cited Able, K. W., D. C. Twitchell, C. G. Grimes, and R. S. Jones. 1987. Sidescan sonar as a tool for detection of demersal fish habitats. Fish. Bull. 85:725-744. Alexander, C. R., and A. M. Simoneau. 1999. Spatial variability in sedimentary processes on the Eel continental slope. Mar. Geology 154:243-254. Auster, P. J., and R. W. Langton. 1999. The effects of fishing on fish habitat. In L.R. Benaka, (ed. I, Fish habitats: essential fish habitat and re- habilitation, p. 150-187. Am. Fish. Soc. Symp. 22, Bethesda, MD. Fnedlander et al.: Sidescan-sonar mapping of benthic trawl marks off Eureka, California 799 Auster, P. J., and N. L. Shackell. 1997. Fishery reser\'es. In J. G. Boreman, B. S. Naka- shima, H. W. Powels, J. A. Wilson, and R. L. Kendall (eds.). Northwest Atlantic groundfish; perspectives on a fishery collapse, p 159-166. Ani. Fish. Soc, Bethesda, Maryland. Auster, P. J., R. J. Malatesta, R. W. Langton, L. Watling, P. C. Valentine, C. S. Donaldson, E. W. Langton, A. N. Shepard, and I. G. Babb. 1996. The impacts of mobile fishing gear on seafloor habi- tats in the Gulf of Maine (Northwest Atlantic): implica- tions for conservation offish populations. Rev. Fish. Sci. 4:185-202. Auster, P. J., C. Michalopoulos, R. Robertson, P. C. Valentine, K. Joy, and V. Cross. 1998a. Use of acoustic methods for classification and moni- toring of seafloor habitat complexity: description of approaches. In N. W. P. Munro and J. H. M. Willison (eds.), Linking protected areas with working landscapes, conserv- ing biodiversity, p 186-197. Science and Management of Protected Areas Association. Wolfville, Nova Scotia. Auster, P. J., C. Michalopoulos, P. C. Valentine, and R. J. Malatesta. 1998b. Delineating and monitoring habitat management units in a temperate deep-water marine protected area. In N. W. P. Munro and J. H. M. Willison (eds.). Linking pro- tected areas with working landscapes, conserving biodiversity, p 169-185. Science and Management of Pro- tected Areas Association. Wolfville, Nova Scotia. Boehlert, G. W. 1996. Biodiversity and the sustainability of marine fish- eries. Oceanography 9:28-35. Bohnsack, J. A. 1996. Maintenance and recovery of reef fishery productivity. In N. V. C. Polunin and C. M. Roberts (eds.). Reef fisher- ies, p. 283-313. Chapman and Hall, London. Bohnsack, J. A., and J. S. Ault. 1996. Management strategies to conserve marine biodi- versity. Oceanography 9:73-82. Botsford, L. W., J. C. Castilla, and C. H. Peterson. 1997. Management of fisheries and marine ecosystems. Science (Wash. D.C.) 277:509-514. Bradstock, M., and D. Gordon. 1983. Coral-like bryozoan growths in Tasman Bay, and their protection to conserve commercial fish stocks. N.Z. J. Mar Freshwater Res. 17:159-163. Brylinsky, M., J. Gibson, and D. C. Gordon. 1994. Impacts of flounder trawls on the intertidal habitat and community of the Minas Basin, Bay of Fundy. Can. Fish. Aquat. Sci. 51:650-661. Caddy, J. F. 1973. Underwater observations on tracks of dredges and trawls and some effects of dredging on a scallop ground. J. Fish. Res. Board Can. 30:173-180. Carlson, H. R., and R. R. Straty. 1981. Habitat and nursery grounds of Pacific rockfish, Sebastes spp., in rocky coastal areas of southeastern Alaska. Mar. Fish. Rev. 43:13-19. Carver, G. A. 1987. Late Cenozoic tectonics of the Eel River Basin re- gion, coastal northern California. In H. Schymiczek and R. Suchsland (eds.). Tectonics, sedimentation and evolu- tion of the Eel River Basin and other coastal basins of north- ern California, p. 61-71. San Joaquin Geological Society Misc. Publ. 37. Chavez, P. S . 1984. U.S. Geological Survey mini image processing sys- tem (MIPS). US Geol. Survey Open-File Rep. 84-880, Reston, VA, 12 p. 1986. Processing techniques for digital sonar images from GLORIA. J. Photogram. Engrg. and Remote Sens. 52:1133-1145. Churchill, J. H. 1989. The effect of commercial trawling on sediment resuspension and transport over the Middle Atlantic Bight continental shelf Continental Shelf Res. 9:841-864. Clarke, S. H., Jr. 1992. Geology of the Eel River Basin and adjacent region: implications for late Cenozoic tectonics of the southern Cascadia Subduction Zone and Mendocino Triple Junction. AAPG (American Association of Petroleum Geologists) Bull. 76:199-224. Clarke S. H., Jr., and M. E. Field. 1989. Geologic map of the northern California continental margin. Map 7A in the California continental margin Geo- logic map series, 1:250,000 scale. California Dep. Mines and Geology, Sacramento, CA. Collie, J. S. 1998. Studies in New England of fishing gear on the sea floor. In E. M. Dorsey and J. Pederson (eds.). Effects of fishing gear on the sea floor of New England, p 53-62. Conservation Law Foundation, Boston. Massachusetts. Collie, J. S., G. A. Escanero, and P. C. Valentine. 1997. Effects of bottom fishing on the benthic magafauna of Georges Bank. Mar. Ecol. Prog. Ser. 155:159-172. Connell, J. H., T. P. Hughes, and C. C. Wallace. 1997. A 30-year study of coral abundance, recruitment, and disturbance at several scales in space and time. Ecol. Monogr. 67:461-488. Dayton, P. K., S. F. Thrush, M. T. Agardy, and R. J. Hofman. 1995. Environmental effects of marine fishing. Aquat. Cons. 5:205-232. Deimling, E. A., and W. J. Liss. 1994. Fishery development in the eastern North Pacific: a natural-cultural system perspective, 1888-1976. Fish. Oceanogr. 3:60-77. de Groot, S. J. 1984. The impact of bottom trawling on benthic fauna of the North Sea. Ocean Manage. 9:177-190. Dorn, M., and M. Saunders. 1997. Status of the coastal Pacific whiting stock in the U.S. and Canada in 1997. In Pacific Fishery Management Council. 1997. Appendix: status of the Pacific coast ground- fish fishery through 1997 and recommended biological catches for 1998: stock assessment and fishery evaluation, 84 p. Pacific Fishery Management Council 2130 SW Fifth Avenue, Suite 224, Portland, Oregon 97201. Dorsey, E. M., and J. Pederson (eds.). 1998. Effects of fishing gear on the sea floor of New England. Conservation Law Foundation. Boston, MA, 160 p. Field, M. E., and J. H. Barber Jr. 1993. A submarine landslide associated with shallow seaf- loor gas and gas hydrates off northern California. In W. C. Schwab, H. J. Lee, and D. C. Twichell (eds.). Submarine landslides: selected studies in the U. S. exclusive economic zone, p. 151-157. U.S. Geol. Surv. Bull. B2002. Field, M. E., and A. E. Jennings. 1987. Seafloor gas seeps triggered by a northern Califor- nia earthquake. Mar. Geology 77:39-51. Field, M. E., J. V. Gardner, and D. B. Prior. 1999. Geometry and significance of stacked gullies on the northern California slope. Mar. Geology 154:271-288. 800 Fishery Bulletin 97(4), 1999 Gardner, J. V., D. B. Prior, and M. E. Field. 1999. Humbolt slide: anatomy of a submarine landslide. Mar. Geology 154:323-338. Gislason, H. 1994. Ecosystem effects of fishing activities in the North Sea. Mar Poll. Bull. 29:520-527. Goff, J. A., D. L. Orange, L. A. Mayer, and J. E. Hughes-Clarke. 1999. Detailed investigation of continental shelf morphol- ogy from a high resolution swath sonar survey over the Eel River Basin, northern California. Mar. Geology ^- 154:255-270. Gordon, D. C, Jr., P. Schwinghamer, T. W. Rowell, J. Prens, K. Gilkinson, W. P. Vass, and D. L. McKeown. 1998. Studies in eastern Canada on the impact of mobile fishing gear on benthic habitat and communities. In E. M. Dorsey and J. Pederson (eds.). Effects of fishing gear on the sea floor of New England, p 63-67. Conservation Law Foundation, Boston, Massachusetts. Greenstreet, S. P. R., I. D. Tuck, G. N. Grewar, E. Armstrong, D. G. Reid, and P. J. Wright. 1997. An assessment of the acoustic survey technique, RoxAnn, as a means of mapping seabed habitat. ICES J. Mar Sci. 54:939-959. Hollander, M., and D. A. Wolfe. 1973. Nonparametric statistical methods. .John Wiley and Sons. New York, NY. 503 p. Hutchings, P. 1990. Review of the effects of trawling on macrobenthic epifaunal communities. Aust. J. Mar Freshwater Res. 41:111-120. lanelli, J. N., R. Lauth, and L. D. Jacobson. 1994. Status of the thornyhead (Sebastelobus sp.) resource in 1994. Appendix D m Status of the Pacific coast ground- fish fishery through 1994 and recommended acceptable biological catches for 1995, 58 p. Pacific Fishery Man- agement Council, Portland, OR Jay, C. V. 1996. Distribution of bottom-trawl fish assemblages over the continental shelf and upper slope of the US west coast, 1977-1992. Can. J. Fish. Aquat. Sci. 53:1203-1225. Jones, J. B. 1992. Environmental impact of trawling on the seabed: a review. N.Z. J. M. Freshwater Res. 26:59-67. Kaiser, M. J., and K. Ramsay. 1997. Opportunistic feeding by dabs within areas of trawl disturbance: possible implications for increased survival. Mar Ecol. Prog. Ser 152:307-310. Kaiser, M. J., and B. E. Spencer. 1994. Fish scavenging behaviour in recently trawled areas. Mar Ecol. Prog. Ser. 112:41-49. Karp, W. A. (ed.). 1990. Developments in fisheries acoustics. Rapp. P.-V. des Reun. 189, 442 p. Lindeboom, H. J., and S. J. de Groot (eds.). 1998. Impact II. The effects of different types of fishing on the North Sea and Irish Sea benthic ecosy.stem. Nether- lands Institute for Sea Research, Den Burg, Texel, The Netherlands, 404 p. Ludwig, J. A., and J. F. Reynolds. 1988. .Statistical ecology. .John Wiley and Sons, New York, NY. 337 p. Matthews, K. R., and L. R. Richards. 1991. Rockfish (Scorpaenidael assemblages of trawable and untrawable habitats off Vancouver Island, British Colum- bia. N. Am. J. Fish. Manage. 11:312-318. Mayer, L. A., J. Huges-Clarke, and S. Dijkstra. 1997. Multibeam sonar: potential applications for fisher- ies research. In G. W. Boehlert, and J. D. Schumacher, (eds.). Changing oceans and changing fisheries: environ- mental data for fisheries research and management, p. 79- 92. U.S. Dep. Commen, NOAA Tech. Memo. NOAA-TM- NMFS-SWFSC-239. Messieh, S. N., T. W. Rowell, D. L. Peer, and P. J. Cranford. 1991. The effects of trawling, dredging and ocean dumping on the eastern Canadian shelf seabed. Continental Shelf Res. 11:1237-1263. NOAA (National Oceanic and Atmospheric Administration). 1996. Our living oceans: report on the status of U.S. living marine resources, 1995. U.S. Dep. Commer., NOAA Tech. Memo, NMFS-F/SPO-19, Washington, D.C., 160 p. Nittrouer, C. A., 1999. STRATAFORM: overview of its design and synthesis of its results. Mar Geology 154:3-12. Nittrouer, C. A., and J. H. Kravitz. 1996. STRATAFORM: a program to study the creation and interpretation of sedimentary strata on continental mar- gins. Oceanography 9:146-152. Pilskaln, C. H., J. H. Churchill, and L. M. Mayer. 1998. Resuspension of sediment by bottom trawling in the Gulf of Maine and potential geochemical consequences. Conserv. Biol. 12:1223-1229. Ramsay, K., M. J. Kaiser, P. G. Moore, and R. N. Hughes. 1997. Consumption of fisheries discards by benthic scav- engers: utilization of energy subsidies in different marine habitats. J. Anim. Ecol. 66:884-896. Reise, K. 1982. Long-term changes in the macrobenthic invertebrate fauna of the Wadden Sea: are polychaetes about to take over? Netherlands .J. Sea Res. 16:29-36. Rogers, S. I., M. J. Kaiser, and S. Jennings. 1998. Ecosystem effects of demersal fishing: a Euporean perspective. In E. M. Dorsey and J. Pederson (eds.). Effects of fishing gear on the sea fioor of New England, p 68- 78. Conser\'ation Law Foundation, Boston, Massachusetts. Rogers, J., L. Jacobson, R. Lauth, J. lanelli, and M. Wilkins. 1997. Status of the thornyhead {Sebastolobus sp.) resource in 1997. Appendix D in Status of the Pacific coast ground- fish fishery through 1997 and recommended biological catches for 1998: stock assessment and fishery evaluation, p. 1-58. Pacific Fishery Management Council, 2130 SW Fifth Avenue, Suite 224, Portland, Oregon 97201. Sainsbury, K.J. 1987. Assessment and management of the demersal fish- ery on the continental shelf of northwestern Australia. In J. J. Polovina and S. Ralston, (eds. ), Tropical snappers and groupers: biology and fisheries management, p. 465- 503. Westview Press, Boulder, Colorado. 1988. The ecological basis of multispecies fisheries and management of a demersal fishery in tropical Aus- tralia. In J. A. Gulland (ed.). Fish population dynamics, 2"'' ed., p. 349-382. John Wiley and Sons, London. Schmitten, R. A. 1996. National Marine Fisheries Service: seeking partners for its National Habitat Plan and identifying essential fish habitats. Fisheries 21:4. Schwinghamer, P., J. Y. Guigne, and W. C. Siu. 1996. Quantifying the impact of trawling on benthic habi- tat structure using high resolution acoustics and chaos theory Can. J. Fish. Aquat. Sci. 53:288-296. Friedlander et al.: Sldescan-sonar mapping of benthic trawl marks off Eureka, California 801 Schwinghamer, P., D. C. Gordon Jr., T. W. Rowcll, J. Prena, D. L. McKeown, G. Sonnichsen, and J. Y. Guignes. 1998. Effects of experimental otter trawling on surficial sediment properties of a sandy-bottom ecosystem on the Grand Banks of Newfoundland. Conserv. Biol. 12: 1215- 1222. Sommerfield, C, and C. N. Nittrouer. 1999. Modern accumulation rates and a sediment budget for the Eel shelf: a flood-dominated depositional environ- ment. Mar Geology 154:227-242. Syvitski, J. P., C. R. Alexander, M. E. Field, J. V. Gardner, D. L. Orange, and J. W. Yun. 1996. Continental-slope sedimentation: the view from northern California. Oceanography 9(3):163-167. Tuck, I. D., S. J. Hall, M. R. Robertson, E. Armstrong, and D.J . Basford. 1998. Effects of physical trawling disturbance in the previ- ously unfished sheltered Scottish sea loch. Mar. Ecol. Prog. Ser. 162:227-242. Veer, H. W., van der, M. J. N. Bergman, and J. J. Beukema. 1985. Dredging activities in the Dutch Wadden Sea: effects on macrobenthic infauna. Netherlands J. Sea Res. 19: 183-190. Watling, L., and E. A. Norse. 1998, Disturbance of the seabed by mobile fishing gear: a comparison to forest clearcutting. Conserv. Biol. 12:1180- 1197. Wiberg, P. L., D. A. Cacchione, R. W. Sternberg, and L. D. Wright. 1996. Linking sediment transport and stratigraphy on the continental shelf Oceanography 9:153-162. Yoklavich, M. M. 1998. Marine Harvest Refugia for West Coast Rockfish. U.S. Dep. Commer, NOAA-Tech. Memo. NMFS-SWFSC- 225. 159 p. Yoklavich, M., R. Starr, J. Steger, H. G. Greene, F. Schwing, and C. Malzone. 1997. Mapping benthic habitats and ocean currents in the vicinity of central California's Big Creek ecological reserve. U.S. Dep. Commer., NOAA-Tech. Memo. NMFS- SWFSC-245, 52 p. Yun, J. W., D.L. Orange, and M. E. Field. 1999. Gas distribution in the Eel River Basin and its link to submarine geomorphology. Mar. Geology 154:357-368. Zar, J. H. 1984. Biostatistical analysis. Prentice-Hall. Inc., Engle- wood Cliffs, NJ, 718 p. 802 Abstract.— The effect of maturation on relative growth of somatic tissues was investigated by measuring and compar- ing monthly changes in dry weight of somatic tissues and reproductive or- gans. In both sexes, reproductive tis- sues grew in relation to total body mass; at maturity female reproductive tissue was 16'7f of total dry body mass, whereas male reproductive tissue was 2.6'7f. In females, the relative mass of thfmantle and head decreased during maturation, whereas the relative mass of the viscera increased. In males, the mass of the viscera increased with maturation, but no decreases occurred. The percentage composition of protein in the mantle and head of females for each maturity stage did not differ sig- nificantly. For both sexes, the digestive gland mass remained relatively con- stant throughout the different maturity stages and seasons, and analysis of stomach fullness indicated that feeding increased in the final maturity stages. All observations support the hypothesis that energy and nutrients for matura- tion are supplied mainly by diet rather than by stored resources, but that dur- ing maturation there is a shift of em- phasis from somatic growth to gonadal development and vitellogenesis. Sepia pharaonis, which appears to be an in- termittent multiple spawner, does not use protein from muscle tissue for de- veloping and growing its reproduction tissues. Reproductive versus somatic tissue growth during the life cycle of the cuttlefish Sepia pharaonis Ehrenberg, 183T Howaida R. Gabr Department of Marine Biology Suez Canal University Ismailia, Egypt Roger T. Hanlon Marine Biological Laboratory Woods Hole, Massachusetts 02543 E-mail address (for R T Hanlon, contact author) rhanlonidimbi edu Salah G. El-Etreby Mahmoud H. Hanafy Department of Marine Science Suez Canal University Ismailia, Egypt Manuscript accepted 16 November 1998. Fish. Bull. 97:802-811 i 19991. Cephalopod mollusks have various life history strategies, ranging from typically semelparous (e.g. many coleoids) to iteroparous (e.g. Nauti- lus spp.X Boyle, 1983, 1987). Differ- ent modes of semelparity may oc- cur, and this diversity may be re- lated in part to growth patterns (Mangold et al., 199.3). Most cepha- lopods that have been studied are fast growing, and they generally reproduce once and die (Calow, 1987). However, others have inter- mittent growth (Boletzky, 1987; Forsythe and Van Heukelem, 1987; Jackson and Choat, 1992) and mul- tiple spawning events (Harman et al., 1989; Lewis and Choat, 1993). We investigated the relation between re- production and somatic growth in the commercially valuable cuttlefish Se- pia pharaonis Ehrenberg, 1831. For many marine invertebrates, reproduction represents an enormous energy investment, and achieving the optimum balance between so- matic and reproductive effort is critical to an individual's lifetime fitness (Calow, 1981). The varia- tions in reproductive strategies usu- ally correlate with variations in feeding and growth patterns. As- sessing the cost of reproduction is difficult, and although slowing of somatic gi'owth may indicate the al- location of energy to oocyte produc- tion, an assessment of somatic tissue may provide a better indicator of the cost of reproduction (Calow, 1983). For example, in some benthic oc- topuses, mantle protein is mobilized to provide energy for egg develop- ment; the subsequent depletion of protein from body tissues probably contributes to the death of some octopus females (Tait, 1986; Pollero and Iribarne, 1988). This would not seem to be a sensible strategy for a swimming cephalopod, such as cuttlefish, which still requires func- tional muscle to survive in the wa- ter column. Sepia pharaonis. which migrates to spawning grounds from its feeding grounds (Gabr et al., 1998), would certainly need to maintain swimming ability. Elucidation of patterns of energy storage and utilization is important in understanding the interaction between organism and environment ( Clarke et al., 1994 ). Although many studies indicate that oocyte produc- Gabr et al.: Reproductive versus somatic growth during the life cycle of Sepia pharaonis 803 tion affects the growth and condition of somatic tis- sue in squid (e.g. Rowe and Mangold-Wirz, 1975; Hatfield et al., 1992), few studies have examined the expenditure of nutrients and energy for the develop- ment of reproductive organs in cuttlefish. Boucaud- Camou (1971) found that the digestive gland of Se- pia officinalis has high concentrations of lipids, sug- gesting that it acts as a storage organ. Castro et al. (1991) studied changes in the digestive gland of .S. officinalis and S. elegans throughout their life cycles and observed that the digestive gland weight of S. officinalis decreased progi'essively with stai-vation. However, there is no evidence that the digestive gland provides stored energy for gonad development. Sepia pharaonis is widely distributed in the Indo- Pacific from the Red Sea to Japan and Australia. Together with S. dollfusi, it is the primary fishery in the Suez Canal and the most valuable commercial cephalopod in the northern Indian Ocean (Nesis, 1987). Little is known about the biology of this spe- cies (Silas et al., 1985; Aoyama and Nguyen, 1989; Gabr et al., 1998). In this paper, we examine aspects of its life history to determine if a trade-off exists between somatic and reproductive tissues. We mea- sured food intake to determine whether nutritional activity was involved in the development of repro- ductive organs. Materials and methods Samples of Sepia phai-aonis were collected monthly from Bitter Lake, the main fishing port in the Suez Canal, from September 1994 to April 1996. In total, 1428 females and 1151 males were examined. The dorsal mantle lengths ranged from 10 to 240 mm. Specimens were dissected to determine sex and ma- turity stage. Four maturity stages for each sex were determined by using a modification of the scale pro- posed by Mangold-Wirz ( 1963) and refined in detail for Sepia pharaonis by Gabr et al. ( 1998 ): for females, I=immature, II=maturing, III=pre-spawning, and IV=spawning; for males, I=immature, II=maturing, III=fully mature, and IV=spawning. Dorsal mantle length (ML) and nidamental gland length (NGL) of females were measured to the near- est mm. The following measurements of mass were made to two decimal places (in grams): total body mass (BM); total somatic mass (SM); mantle mass (MM); head mass including arms and tentacles (HM); digestive gland mass (DGM); viscera mass (VM) in- cluding gills, stomach, caecum, pancreas, ink sac; ovary mass ( OM ) including oviduct; testis mass ( TM ); nidamental gland mass (NGM); and spermatophoric complex mass (SCM). Relative assessment of somatic tissue growth Two statistical analyses were conducted to determine if somatic tissue declined in relation to size during maturation. All statistical analyses were carried out by using the MINITAB statistical package and all data were logjg-transformed. In the first analysis, a multiple regression was applied to the female data only by using logjg mantle length and logjQ nidamental gland length as inde- pendent variables. These variables were selected because they correlated with our maturity scale (Gabr et al., 1998). Multiple regression equations were obtained for log-transformed total somatic mass and for masses of mantle, head, digestive gland, vis- cera, and ovary (including oviduct and oviductal gland, and nidamental gland). This analysis allowed for the use of a continuous variable in the assess- ment of maturity over the whole size range of the sampled population. This multiple regression analy- sis was carried out only on females because length was used in the analysis. To carry out a similar exer- cise with male cuttlefish, the relation of the log-trans- formed mass of each organ with spermatophoric com- plex mass (SCM) and body mass would have to be used. These variables would lead to problems in scal- ing because they would involve autocorrelation, which would distort perceived relationships (LaBar- bera, 1989). The ratio of SCM to ML has not been tested before as an objective measure of maturity in male cuttlefish and therefore no analysis was in- cluded here. In the second analysis, a multivariate analysis of covariance (ANCOVA) was applied to the total so- matic mass, individual somatic organs, and repro- ductive organs, whose relationships with the repro- ductive cycle were considered dependent variables (Garcia-Berthou and Moreno-Amich, 1993). Mantle length was ti'eated as the covariate. A fundamental assumption of standard ANCOVA (McCullagh and Nelder, 1983) is the homogeneity of regression coef- ficients (slopes) of dependent variable and covariate relationships. This assumption can be tested with a special design of ANCOVA, by analyzing the pooled covariate by factor interaction. If the covariate by factor interaction is significant, standard ANCOVA should not be developed. Otherwise, if the covariant by factor interaction is not significant, the standard ANCOVA design is the preferred method. For the cases with significant effect with factor (maturity stage), the variation can be described by using the predicted means for each cell, adjusted for the effect of the covariate. We set the alpha level for statistical significance at 0.001 to identify the strongest effects of maturity on the dependent variables. 804 Fishery Bulletin 97(4), 1999 Dry weight analysis M m 3 n 20 To study the seasonal variation in somatic and reproductive organs, dry weight was used to correct for differences in water content be- tween different organs. The samples for this study were collected from November 1994 to April 1996. From each monthly sample, a subsample of 20 females and 20 males was selected with mantle lengths within a 20-mm size range encompassing the mean size at maturity (the size at which 50% of the individuals were maturing or mature). This size was 122 and 61 mm ML ( calculated by using the whole-year sample) for females and males, respectively (Gabr et al., 1998). It is possible to treat speci- mens from around the mean as "standard" animals (sensii Gabbott, 1976) and so com- pare cuttlefish at different degrees of maturity with- out having to make statistical coiTections for size. There were no cuttlefish of maturity stage I in this sub- sample; however, stages II to IV were present. The cuttlefish in the subsample were dissected in the same way as those in the sample as a whole. The same measurements of mass to two decimal places (in grams) were taken, but dry rather than wet weight was used. Tissues were dried to constant weight (for 48 h) at 80°C and then cooled in a desiccator The following indices were calculated for dry weight tissues; gonadosomatic index for females (GSI) - 100 OM/BM; gonadosomatic index for males (GSI) = 100 TM/BM: nidamental gland index (NGI) - 100 NGM/BM; spermatophoric complex index (SCI) = 100 SCM/BM; mantle index (MI) = 100 MM/BM; head index (HI) = 100 HM/BM; digestive gland in- dex ( DGI ) = 100 DGM / BM, and viscera index ( VI ) = 100 VM/BM. Analysis of protein The dry tissue of mantle and whole head for 7 to 9 females of maturity stages II-IV was ground to a fine powder with a mortar and pestle. The powdered samples were placed in sealed plastic vials and stored for further analysis. The Kjeldahl method for deter- mination of nitrogen was used (see Giese, 1967). % Protein = % nitrogen x 6.25 (Giese, 1967; Rigby, 1990; Dickey-Collas, 1991). Kruskal-Wallis ANOVA was used to examine the significant variation in per- centage of protein between maturity stages. Female Male mmm aDD D OD D D iititm wi ri I (niDCKD O IDCO 0.05). As expected, there was also a highly significant increase (P<0.05) in repro- ductive organ mass with maturity (NGL). The second analysis was an analysis of multivari- ate covariance (ANCOVA). In an animal that matures at a wide range of body sizes, it is difficult to quan- tify the pattern of maturation because comparisons of individuals of different sizes cannot be thought of as indicating the pattern of growth and maturation of any one animal. Thus, although some animals matured at 60 mm ML, others remained immature at larger sizes ( Fig. 1). ANCOVA corrects for the varia- tion in size (ML) of animals within maturity stages. Table 2 shows the initial standard ANCOVA de- sign to test for homogeneity of slopes. In females, there was significant effect of an interaction between the factors (maturity stage) and covariate (mantle length) for OM and NGM. The effect of interaction of mantle length and maturity stages on TSM, MM, HM, DGM, and VM was not significant (P>0.001). Thus in these cases the hypothesis of homogeneity of slopes was accepted; therefore further analyses were required to test the effect of maturity stages for these variables. In males, there was a highly sig- nificant effect (P<0. 001) between mantle length and maturity stage for all log-transformed dependent variables except spermatophoric complex mass. Thus for these variables the hypothesis of homogeneity of slopes was rejected and further analysis was not pos- sible (i.e. the standard ANCOVA design should not be developed for these cases). The results of the final ANCOVA design (in the cases where the effect of mantle length and matu- rity stage interactions were not significant) are indi- cated in Table 3. In female Sepia pharaonis, matu- rity significantly affected TSM, MM, HM, and VM (P<0.001) but did not significantly affect DGM (P>0.001 ). In males, maturity significantly influenced spermatophoric complex mass (P<0.001). Figure 2 shows the adjusted means for each de- pendent variable for each maturity stage after the removal of the mantle length (covariate) effect when the effects of maturity stages were found to be sig- nificant. For females, there was a decrease in total somatic mass, mantle mass, and head mass with in- creasing maturity stage. There was a decrease in viscera mass from stage I to stage III followed by an increase at stage IV (Fig. 2A). For males, there was an increase in relative mass of spermatophoric com- plex mass with increasing maturity stage (Fig. 2B). Monthly variation in soma versus gonad production Monthly fluctuations in mean dry weight indices for individual organs of somatic tissue (mantle index, MI; head index, HI; digestive index, DGI; and vis- cera index, VI) and reproductive tissue (gonadoso- matic index, GSI; nidamental gland index, NGI; and spermatophoric complex index, SCI) for females and males are illustrated in Figures 3 and 4, respectively. These indices displayed monthly fluctuations and indicated the relative proportion that each body com- ponent contributed to the weight of the entire body. The mantle mass was always the largest component 806 Fishery Bulletin 97(4), 1999 Table 2 Preliminary design of ANCOVA for Sep a pharaonis. In each case, mantle length is the covariate maturity stage I-IV) is the factor, and the dependent variables are total somatic mas s (TSM), mantle mass (MM), head mass (HM), digestive gland mass (DGM), viscera mass (VM), ovary mass (OM), nidamental gland mass (NGM), testis mass (TM), and spermatoph aric complex mass (SCMl The homogeneity of slopes was tested with a poc led covariate by factor interaction: F-statistic. Signif cance levels (P<0.001) are indicated by an asterisk. w^ Mantle length (ML) F P Maturity stages (MSt) F P ML xMSt F P Female TSM 11000.00 <0.001* 2.80 0.039 1.87 0.130 MM 11000.00 <0.001* 4.74 0.003 2.81 0.030 HM 10000.00 <0.001* 5.29 0.001 3.93 0.009 DGM 3440.00 <0.001* 2.18 0.089 2.16 0.092 VM 2926.30 <0.001* 0.40 0.754 0.98 0.400 OM 271.07 <0.001* 25.15 <0.001* 29.87 <0.001* NGM 513.73 <0.001* 166.10 <0.001* 317.58 <0.001* Male TSM 918.60 <0.001* 22.00 <0.001* 23.00 <0.001* MM 1745.94 <0.001* 14.32 <0.001* 13.66 <0.001* HM 1295.97 <0.001* 12.50 <0.001* 11.02 <0.001* DGM 849.18 <0.001* 11.15 <0.001* 9.55 <0.001* VM 853.13 <0.001* 5.96 <0.001* 5.28 <0.001* TM 612.65 <0.001* 294.18 <0.001* 15.52 <0.001* SCM 749.19 <0.001* 6.01 <0.001* 0.88 0.450 proportionally, followed by the head mass. Thus, S. pharaonis weight gain and loss were accounted for mainly by the mantle and head mass that underwent fluctuation in mean dry weight index during the study. The MI and HI diminished considerably as gonad development proceeded. The largest decrease in Ml and HI of females occurred in spring and summer (from March to September), as shown in Figure 3. These indices decreased from 52% to 41% and from 34% to 28%, respectively. In contrast, during the same period the gonads (GSI, NGI ) exhibited devel- opment; GSI increased from 1% to 9% and NGI in- creased from 2% to 7%. VI increased rapidly from May to November, from 5% to lO'/f . DGI showed little change over the whole sampling period (ranging be- tween 7% and 10% ). By contrast, in male Sepia pharaonis, the MI, HI, and DGI remained relatively constant throughout the entire sampling period (Fig. 4). The pattern of VI was consistent with the situation in female, increas- ing from 5% to 8% during the period from May to October There was a gradual increase in GSI and SCI from November, reaching the highest value in May (=1.3%), and spawning after June was indicated by a decline in GSI and SCI. Table 3 Final des ign of ANCOVA for Sepia pharaonis data for which the homogeneity hypoth esis was accepted, F-statistics, and P-values See Table 2 for definitions of abbreviations. Big- | nificance levels (P<0.001) are in dicated by an asterisk. Mantle length (ML) Maturity stages (MSt) F P F P Female TSM 15000.00 <0.001* 25.130 <0.001* MM 16000.00 <0.001* 40.570 <0.001* HM 14000.00 <0.001* 10.470 <0.001* DGM 4770.00 <0.001' 0.680 0.560 VM 4031.89 <0.001* 11.270 <0.001* Male SCM 901.800 <0.001' 133.890 <0.001' Percentage of protein, water content, and maturity stage The mean percentage composition of protein and mean percentage of water content in the mantle and head of females for maturity stages II-IV is shown Gabr et al : Reproductive versus somatic growtli during the life cycle of Sepia pharaonis 807 A 2.27 1 .82 S 2.25 Female TSM 1.80 Female HM E 2.23 J\ 5 1-79 I\ 1 2.21 0 2 2.19 ^vr 1 1.78 \rr o \ "" 1 .75 X^ 9 1*^ * -ri- £.13 1 ./U 1.94, 1,16 [ S 192 X Female MM X Female VM 1 1.90 ^^"\-r 1 1 .08 \ log mantle GD CD en 00 log viscera i 8 i N^ 1.84 * ■ ■ n rtc L 1 II III IV 1 II III IV Maturity stage Matunty stage B X 0.00 , ■E. -0.10 Male SCM S -0.20 ^^ ■c M, -0.30 •i 1 -0.40 2 e 1 -0.50 ^ ^ -0.60 li^ M -0.70 1 II III IV Matunty stage Figure 2 (A) Adjusted means (SD) for dependent variables (total somatic mass, TSM; mantle | mass, MM; head mass, HM; viscera mass.VM 1 at each maturity stage for female Sepia pharaonis. (B) Adjusted means (SD) for dependent variable spermatophoric complex | mass (SCM) for males. Adjusted means (SD) were obt ained by analysis of covariance removing the effects of the covariate (mantle length!. in Table 4. In neither mantle nor head was there sig- nificant variation (P>0.05) in percentage of protein or water content with maturity stage. Stomach fullness in relation to maturity stages The stomach fullness for each maturity stage for each sex is illustrated in Figure 5. The degree of stomach fullness remained relatively constant through the maturity stages, indicating that Sepia pharaonis continued to feed when mature. Length-weight relationship The length-weight relationship for Sepia pharaonis is described by the equation: Wt = 0.27L'~^-' , r- = 0.99, n = 966 for females, and Wt = 0.28L2 ^o , ,-2 = o.99, n = 723 for males. The elevation and slopes from the re- gression equations for females and males did not dif- fer significantly, thus indicating that females have approximately the same weight as males at the same mantle length, and that both sexes increase simi- larly in weight per unit gain in mantle length. 808 Fishery Bulletin 97(4), 1999 Discussion This study describes the pattern of allocation of re- sources to the growth of somatic and reproductive tissues in Sepia pharaonis, one of the two most com- mercially important cephalopods in the Suez Canal and northern Indian Ocean. This study indicates that gametogenesis is fueled by energy and nutrients de- rived from the diet rather than from reallocation of somatic reserves. Two complementary methods (analysis of covariance and multivariate regression on the entire data set) and the monthly changes in 60 50 40 30 20 I i i I N94 J M M J S N95 J M 12 N94 M M N95 Figure 3 Female Sepia pharaonis mean (SD) dry weight indices of mantle (MI) head (HI), viscera (VI), digestive gland (DGI), nidamental gland (NGI) and gonad masses (GSI) for standard size (110-1.30 mm MLi. dry weight of somatic indices of a restricted size range showed essentially similar patterns of maturation. Sexual maturation in Sepia pharaonis affected males and females differently. Forsythe and Van Heukelem (1987) suggested that male maturity in cephalopods was achieved at little cost to somatic growth and that, in most cases, males continue to grow after reaching maturity. Changes in male so- matic mass do not appear to be related to maturity. At full maturity in males, dry weight of reproduc- tive and accessory reproductive tissues accounted for only 2.6% of the total somatic dry weight. However, these tissues accounted for approxi- mately 16*^ in females. For both sexes, somatic growth was completed by late October as the average mantle mass and head mass components reached maxi- mum weights. In females, spawning occurred from spring throughout the summer, and the end of spawning activity was reflected by a sharp drop in the average gonad weight index in October (Fig. 3) . Thus spawning occurred at the time of in- creasing temperature and food avail- ability (Gabr, unpubl. data). There was no evidence of decline in feeding rate with maturation. This finding suggests that gonadal growth may be supported directly by the assimilated food ration, although some diversion of resources into reproduction at the expense of so- matic growth also occurred. This diver- sion of resources was indicated by a small but significant decrease in mantle and head mass with maturation (Fig. 2), and the inverse relationship of mantle . and head mass with gonadal indices (Fig. 3). There is some evidence to support the idea that female Sepia pharaonis may feed, grow, and mature simultaneously in the field. For example, no spent fe- males were caught at any time of the year. This may be because it is difficult to differentiate spent females from ma- ture ones if spent females remain in good condition (because they continue to feed when mature). Because females need a greater energetic reserve for re- production, they grow faster than males and reach a maximum size of 240 mm ML, whereas the maximum size for males was 170 mm ML. Sepia pharaonis spawns in shallow water and migrates M Gabr et al Reproductive versus somatic growth during the life cycle of Sepia pharaonis 809 60 50 40 30 20 N94 r 12 N94 offshore after spawning either to die or to return to spawn again; such a migration is un- hkely to be preceded by any deg- radation of mantle muscle, which is the major organ for lo- comotion in cuttlefish. In nei- ther mantle nor head was there a significant variation in per- centage of protein or water con- tent with maturity stage. These data indicate that female S. pharaonis possibly do not use mantle or head tissue to fuel gonadal development, at least during the beginning of the spawning season. In sharp contrast, studies of another cuttlefish. Sepia doll- fusi (Gabr et al., in press), have demonstrated that this species seems to use mantle muscle as an energy source during go- nadal development. There thus appears to be a continuum in cuttlefish reproductive strategy. At one end, spawning is associ- ated with gonad maturation at the expense of somatic tissue. At the other end, growth of re- productive organs takes place through the use of energy and nutrients derived from the diet, not through mobilization of nu- trients and energy from somatic tissue. The results of this paper agree with some studies on the squid lUex argentinus (Hatfield et al., 1992; Rodhouse and Hatfield, 1992; Clarke et al., 1994). Although mantle mass of female lUex argentinus decreases in relation to ML with matu- rity, this is not associated with degradation or changes in the biochemical composition of mantle muscle. Studies on Lo//^o /brfoesj (Collins etal., 1995), L. gahi (Guerra and Castro, 1994) and Photololigo sp. (Moltschaniwskyj, 1995) have also demonstrated that maturation and growth occur simultaneously during most of the life cycle and that the condition of the squid remains high at maturation. The digestive gland showed no loss of mass during maturation. Boucaud-Camou (1971) indicated that the digestive gland of Sepia officinalis could act as a lipid storage organ, and Blanchier and Boucaud- MI HI i 5- i i 5 I M M N95 VI DGI M M N95 M 24 1.6 0.8 0.0 SCI GSI N94 M N95 M Month Figure 4 Male Sepia pharaonis mean (SD) dry weight indices of mantle (MI), head (HI), vis- cera (VI), digestive gland (DGI), spermatophoric complex (SCI) and gonad masses (GSI) for standard size (50-70 mm ML). Camou (1984) found that the variation in lipid lev- els of the digestive gland seemed more related to diet than to sex or maturity state of the gonads. There was no significant variation in the digestive gland weight with maturity state and season. Thus, the digestive gland mass cannot be used as an indicator of feeding activity in S. pharaonis as it can in some squid (Abolmasova et al., 1990) because no increase in DGI was found in the season of high feeding ac- tivity. This agi-ees with Castro et al. ( 1991 ), who found no significant variation in the digestive gland weight with season in S. officinalis and S. elegans. In conclusion, this study shows that the pattern of intermittent spawning suggested by our previous 810 Fishery Bulletin 97(4), 1999 Table 4 Summary of mean ± SD values of percentage of protein and water content in mantle and head dry tissue with maturity stages for female Sepia pharaonis. % Protein % Water content Maturity stage Number Mantle Head Number Mantle Head II 9 78.33 ±0.68 76.21 ±0.32 66 75.83 ±1.4 79.38 ±1.4 III 8 79,78 ±0.41 76.23 ±0.39 20 76.18 ±1.3 79.64 ±1.06 IV 7 78.19 ±0.39 75.82 ±0.53 79 75.48 ±1.8 78.91 ±1.5 Kruskal-Wallis ANOVA: H 2.08 0.24 1.32 3.37 P 0.35 0.88 0.51 0.18 100 o study of this species (Gabr, et al., 1998) is quite likely, partly because the spawning season is greatly protracted (Fig. 3). Indeed, it appears that this species is capable of meeting the demands of oo- cyte production without me- tabolizing reserves of other tissues. Both males and fe- males continue to feed as ma- turity is reached, thus energy for gonad production is di- verted primarily from the food supply, although some diversion from somatic tissue may also occur. In the context of this paper and results from S. doll fust (Gabr et al., in press), variation in reproduc- tive strategies correlates with variations in feeding and growth patterns as suggested previously for other coleoids (Mangold et al., 1993). Acknowledgments We are especially grateful to Emma Hatfield for her help and advice about data interpretation during preparation of the manuscript. We also wish to ex- press our thanks to Michael Maxwell, Paul Rago, Jean Boal, and Nadav Shashar for frequent advice and assistance in statistical analysis. This study, part of a doctoral dissertation, was supported by the Em- bassy of the Arab Republic of Egypt, Cultural and Educational Bureau. A portion of this study was sup- ported by the British Council through Liverpool University and the Marine Biology Department, Suez Canal University. Female Male 100 o I u in Maturity stages I II III IV Maturity stages Figure S Stomach fullness for each maturity stage of both sexes of Sepia pharaonis. 0 = empty; 1 = one-quarter full: 2 = half full; 3 = three-quarters full; 4 = full or distended). Literature cited Abolmasova, G. I., G. E. Shul'man, A. M. Shchepkina, and G. F. Dziganshin. 1990. Content of dry matter in the liver of the squid Sihenoteuthis pteropus in the eastern Atlantic Ocean as an index of trophicity. Oceanology 30:359-362. Aoyama, T., and T. Nguyen. 1989. Stock assessment of cuttlefish off the coast of the people's democratic republic of Yemen. Shimonoseki Univ. Fisheries. 37 (2.31:61-112. Blanchier, B., and E. Boucaud-Camou. 1984. Lipids in the digestive gland and the gonad of imma- ture and mature Sepia officinalis (Mollusca: Cephalopoda I. Mar Biol. 80:39-43. Boletzky, S.v. 1987. Fecundity variation in relation to intermittent or chronic spawning in the cuttlefish. Sepia officinalis L. (Mollusca. Cephalopoda). Bull. Mar Sci. 40i2):382-387. Boucaud-Camou, E. 1971. Constituants lipidiques du foie de Sepia officinalis. Mar Biol. 18:66-69. Gabr et al,; Reproductive versus somatic growth during the life cycle of Septa pharaonis 811 Boucher-Rodoni, R., E. Boucaud-Camou, and K. Mangold. 1987. Feeding and digestion. //! P. R. Boyle led.), Cepha- lopod life cycles, vol. 2, p. 85-108. Academic Press, Lon- don, England. Boyle, P. R. (ed.). 1983. Cephalopod life cycles, vol. 1: species accounts. Aca- demic Press. London, England, 475 p. 1987. Cephalopod life cycles, vol, 2: comparative reviews. Academic Press, London, England, 441 p. Calow, P. 1981. Resource utilization and reproduction. In C. R. Townsend (ed.). Physiological ecology, p. 245-270. 1983. Life-cycle patterns and evolution. In W. D. Russell- Hunter (ed.). The Mollusca, vol. 6, p 649-678. Academic Press, New York, NY. 1987. Fact and theory- an overview. In P. R. Boyle (ed.), Cephalopod life cycles, vol. 2, p. 351-365. Academic Press, London. England. Castro, B. G., A. Guerra, and C. M. F. Jardon. 1991. Variation in digestive gland weight ofSepia officinalis and Sepia etegans through their life cycles. //; E. Boucaud- Camou, (ed.). Acta of the first international symposium on the cuttlefish Sepia, p. 99-103. Centre de Publications de rUniversite de Caen. Clarke, A., P. G. Rodhouse, and D. J. Gore. 1994. Biochemical composition in relation to the energet- ics of growth and sexual maturation in the ommastrephid squid lUexargentinus. Phil. Trans. R. Soc. Lond. 344:201- 212. Collins, M. A., G. M. Burnell, and P. G. Rodhouse. 1995. Reproductive strategies of male and female Loligo forbesi (Cephalopoda: Loliginidae). J. Mar. Biol. Assoc. U.K. 75:621-634. Dickey-Collas, M. 1991. Studies on the effect of enriched food on the growth performance and pigmentation of flatfish larvae. Ph.D. diss., LIniv. Liverpool. LI.K. Forsythe, J. F., and W. F. Van Heukelem. 1987. Growth. In PR. Boyle ( ed. ), Cephalopod life cycles, vol. 2. p 135-156. Academic Press, London, England. Gabbott, P. A. 1976. Energy metabolism in marine mussels. In B. L. Bayne, (ed.), Marine mussels: their ecology and physiol- ogy, p 293-355, Cambridge LIniv. Press, Cambridge. Gabr, H. R., R. T. Hanlon, M. H. Hanafy, and S. G. El-Etreby. 1998. Maturation, fecundity and seasonality of reproduc- tion of two commercially valuable cuttlefish. Sepia pharaonis and S. doUfusi. in the Suez Canal. Fish. Res. 36:99-115. In press. Reproductive versus somatic tissue allocation in the cuttlefish SepjQ dollfusi Adam (1941). Bull. Mar Sci. Gareia-Berthou, E., and R. Moreno-Amich. 1993. Multivariate analysis of covariance in morphomet- ric studies of the reproductive cycle. Can. J. Fish. Aquat. Sci. 50:1394-1399. Giese, A. C. 1967. Some methods for the study of the biochemical con- stitution of marine invertebrates. Ocean. Mar. Biol. Ann. Rev. 5:159-186. Guerra, A., and B. G. Castro. 1994. Reproductive-somatic relationships in Loligo gahi (Cephalopoda: Loliginidae) from the Falkland Islands. Antarct. Sci. 6:175-178. Harman, R. F., R. E. Young, S. B. Reid, K. Mangold, T. Suzuki, and R.F. Hixon. 1989. Evidence for multiple spawning in the tropical oce- anic squid Sthenoteuthis oualaniensis (Teuthoidea: Omma- strephidae). Mar. Biol. 101:513-519. Hatfield, E. M. C, P. G. Rodhouse, and D. L. Barber. 1992. Production of soma and gonad in maturing female Illex argentinus (Mollusca: Cephalopoda). J. Mar. Biol. Ass. U.K. 72:281-291. Jackson, G. D., and J. H. Choat. 1992. Growth in tropical cephalopods: an analysis based on statolith microstructure. Can. J. Fish. Aquat. Sci. 49 (2):218-228. LaBarbera, M. 1989. Analyzing body size as a factor in ecology and evolution. A. Rev. Ecol. Syst. 20:97-117. Lewis A. R., and J. H. Choat. 1993. Spawning mode and reproductive output of the tropi- cal cephalopod Idiosepius pygmaeus. Can. J. Fish. Aquat. Sci. 50(l):20-28. Mangold-Wirz, K. 1963. Biologie des cephalopodes benthiques et nectoniques de la Mer Catalane. Vie Millie, suppl. 13, 285 p. Mangold, K., R. E. Young, and M. Nixon. 1993. Growth versus maturation in cephalopods. In T Okutani, R. K. O'Dor. and T Kubodera (eds.). Recent ad- vances in cephalopod fisheries biology, p. 697-704. Tokai LTniv. Press, Tokyo, Japan. MeCullagh, P., and J. A. Nelder. 1983. Generalized linear models. Chapman and Hall. London, England, 261 p. Moltschaniwskyj, N. A. 1993. Multiple spawning in the tropical squid Photololigo sp.: what is the cost in somatic growth? Mar. Biol. 124(1 ):127-135. Nesis, K.N. 1987. Cephalopods of the world. TFH. Publications, Nep- tune City NJ, 351 p. Pollero, R. J., and O. O. Iribarne. 1988. Biochemical changes during the reproductive cycle of the small Patagonian octopus. Octopus tehuelchus, dVrb. Comp. Biochem. Physiol. (B) 90:317-320. Rigby, M. J. 1990. Studies on the rearing of larval and post larval tur- bot iScophthalmus maximus) using enriched live foods, with special emphasis on fatty acids. Ph.D. diss., Univ. Liverpool. U.K. Rodhouse, P. G., and E. M. C. Hatfield. 1992. Production of soma and gonad in maturing male Illex argentinus (Mollusca: Cephalopoda). J. Mar. Biol. Assoc. U.K. 72:293-300. Rowe, V. L., and K. Mangold-Wirz. 1975. The effect of starvation on sexual maturation in Illex illecebrosus (Lesueur) (Cephalopoda: Teuthoidea). J.Exp. Mar. Biol. Ecol. 17:157-164. Silas, E. G., R. Sarvesan, K. P. Nair, Y. A. Sastri, P. V. Sreenivasan, M. M. Meiyappan, K. Vidyasagar, K. S. Rao, and B. N. Rao. 1985. Some aspects of the biology of cuttlefishes. Cephalo- pod bionomics, fisheries and resources of the exclusive eco- nomic zone of India. Cmfri. Bull. 7:49-70. Tait, R.W. 1986. Aspects physiologiques de la senescence post-repro- ductive chez Octopus vulgaris. Ph.D. diss., L'Lfniversite Paris VI, Pans, 250 p. 812 Abstract.— Dorsal spines, otoliths, scales, and vertebrae collected from yellowtail kingfish (Seriola lalandi) in NSW, Australia, were assessed for use- fulness in estimating age. Legibility of growth zones, the time scale at which zones form, and precision of age esti- mates were evaluated for fish sizes from 323 to 1090 mm FL. All calcified structures contained growth zones, but dorsal spines were unsuitable for age determination because it was likely that early growth zones were lost. From marginal increment analysis, it ap- peared that one zone was laid down per year for otoliths and possibly scales, but a clear pattern was not found for ver- tebrae. Although exact agreement be- tween repeated age readings was rela- tively low (50-669!-), agreement within one zone was higher (92-96'7f) and scales provided the most precise read- ings. Precision decreased with increas- ing age of fish. Growth curves derived from otoliths and scales were similar for all ages except fish from the first age class; those derived from otoliths and vertebrae were similar for all fish with less than eight growth zones. Al- though statistical differences were found between the growth curves of scales and vertebrae for some age classes, with the exception of the first age class these differences were not bio- logically important. Growth rates esti- mated from length-frequency (age- based) and mark-recapture (length- based) data compared favorably with those estimated from calcified aging structures. Otoliths, scales, and verte- brae all showed promise as structures for aging kingfish, but further work is needed to determine the position of the first zone and to validate estimates for all age classes. Until such work is com- pleted, we recommend that scales and either otoliths or vertebrae be used for aging kingfish. Aging methods for yellowtail kingfish, Seriola lalandi, and results from age- and size-based growth models Bronwyn M. Gillanders NSW Fisheries Research Institute PO Box 21, Cronulla, NSW 2230, Australia Present address: School of Biological Sciences A08 University of Sydney New South Wales 2006, Australia E-mail address bronwynia'bio usyd edu au Douglas J. Ferrell Neil L. Andrew NSW Fisheries Research Institute PO Box 21 Cronulla, New South Wales 2230, Australia Manuscript accepted 9 December 1998. Fish. Bull. 97:812-827 (1999). Yellowtail kingfish, Seriola lalandi, is a popular recreational species and supports significant commercial fisheries throughout temperate re- gions of the world. In Australia, the major commercial fishery for yel- lowtail kingfish is in New South Wales where 400-600 tonnes are caught per year. Despite the exist- ence of a commercial fishery and persistent controversy about exploi- tation of the species, very little is known about its biology. For future management of the fishery and for stock assessment purposes, informa- tion on age and growth is needed. Estimates of age of Seriola spp. have been derived from a variety of methods and structures, reflecting the difficulty of aging Seriola and other pelagic species. The Japanese species (Seriola quinqueradiata) is the most studied species in the ge- nus because of its importance in aquaculture. Studies in the 1950s used scales, vertebrae, and opercu- lar bones for aging (e.g. Mitani, 1955, 1958; Mitam and Sato, 1959). More recent studies have focused on the use of vertebrae (e.g. Munekiyo et al., 1982; Murayama, 1992). Scales have also been used to age S. lalandi (formerly S. dorsalis. Baxter, 1960), as have otoliths (Penney^). Recently, sectioned oto- liths were used to determine ages ofSeriola dumerili, although it was acknowledged that age and growth determinations were difficult (Man- ooch and Potts, 1997). Despite the use of a variety of structures for ag- ing Seriola spp., there has been no comparative analysis to determine which structure provides the most reliable estimates. In addition, most studies have assumed that growth zones are annual and there have been few validations of age estimates (but see Mitani and Sato, 1959; Baxter, 1960). The specific objectives of this study were 1) to assess the useful- ness of several structures (scales, otoliths, dorsal spines, and verte- brae) for determining the age of kingfish, 2) to compare multiple age estimates for different structures in order to determine the most precise method for determining age and growth parameters, 3) to provide information on size-at-age and 4) to compare growth rates obtained Penney, A. J. 1992. Sea Fisheries Re- search Institute, Private Bag X2, Rogge Bay 8012, South Africa. Personal commun. Gillanders et al.: Aging methods for Serio/a lalandi 813 from age-length data (from counts of zones in calci- fied structures) to those obtained from length-fre- quency and mark-recapture (tagging) data. We ac- knowledge that the data from these three approaches for estimating growth are not directly comparable (Francis, 1988a), but follow the recommendations of Francis (1995) in interpreting differences. Materials and methods Fish collection and treatment Yellowtail kingfish, Seriola lalandi. specimens were collected from New South Wales, Australia, between August 1995 and July 1996 by commercial or recre- ational fishermen. Fish caught by commercial fish- ermen were obtained after being processed by fillet- ing. Fish were measured (total length, fork length) and sagittal otoliths, dorsal spines, scales, and ver- tebrae removed. Dorsal spines were examined but considered unsuitable for aging because the center (core) region was either occupied by vascular bony tissue or was hollow. The hollow core in large fish was found to have a larger diameter than the whole spine of small fish. For this reason, it was likely that early growth zones were lost in older fish and there- fore spines were considered unsuitable for aging. Sagittae Whole sagittae were burned for 7 min at 500'C. They were viewed under a low-power dissecting microscope (6x magnification) with reflected light against a black background. Assignments of age were based on counts of opaque (light) zones or ridges (or both) that were usually most visible at the base of the rostrum on the ventral surface (Fig. lA). Sagittae were also embedded in clear resin, sectioned in a transverse plane with a low speed saw, the sections (=350 mm thickness) mounted on glass slides, and viewed un- der a compound microscope (40x magnification) with reflected light against a black background. Scales Scales were removed from a position anterior and ventral to the pectoral fin. It was necessary to re- move scales from such a position because most fish had been processed prior to the removal of scales. Scales from each fish were soaked in a solution of sodium hydroxide for 3 h, then rinsed and soaked in water for a further 3-12 h. Clean, nonreplacement (i.e. original scales showing typical ctenoid shape) scales were dry-mounted between two glass micro- scope slides. Scales were read under a compound microscope (20x magnification) with reflected light against a black background. Presumed annuli were identified by cutting over {sensu Bagenal and Tesch, 1978) in the lateral fields or by clear zones, where circuli were more widely spaced, in the anterior field. Vertebrae The second vertebra of 24 vertebrae present in king- fish was chosen because it was most easily obtained from processed fish. Vertebrae were either stored fro- zen with flesh intact, or the flesh was removed, and the vertebrae were separated from each other and stored dry. The spines were removed and each verte- bra cut in half along the longitudinal— horizontal plane and stained in a solution of alizarin red S (fol- lowing Berry et al. , 1977 ) for 8 h, rinsed in tap water for at least 1 min, and dried at room temperature. Vertebrae were read under a dissecting microscope (6-12x magnification) with reflected light from a blue-filtered, high-intensity bulb against a black background. Age was estimated from counts of ridges on the inner surface of the vertebra from the core to the outer edge of the centrum (Fig. IC). Assessment of aging techniques To determine whether the zones would be reliably interpreted, two replicate counts of zones were made for each structure by the same person. Counts were usually separated by one month. All readings were done in a random order, with no knowledge of date of collection, size of fish, or knowledge of previous counts. Preliminary investigations of transverse sec- tions of otoliths from 50 fish (ranging from 323 to 1090 mm FL) found that growth zones were not in- terpretable for any fish (Fig. 2). Multiple counts of zones (two counts for each ag- ing structure ) were used to estimate the probability of assigning an age a to a fish with estimated "true" age b following maximum-likelihood estimation pro- cedures outlined in Richards et al. ( 1992). This pro- cedure requires the estimation of a classification matrix where there are columns for each "true" age and rows for each assigned age, and the entries re- fer to probabilities of assigning age a to a fish, given its true age 6. "True" age is best described as the most probable age, and it does not refer to the accu- racy of the age estimate nor does it substitute for age validation procedures. It is assumed that fish will be assigned to the true (or most probable) age class with the highest probability (Richards et al., 1992). The classification matrix is defined by up to four parameters, where the first two parameters, CTj and 814 Fishery Bulletin 97(4), 1999 Figure 1 Structures used to estimate the age of kingfish from New South Wales; (A) otolith i860 mm FL, age 3i; (B) scale (618 mm FL, age 3); and (C) vertebra (720 mm FL, age 3). All structures viewed with reflected light against a black background. Size of scale bars are indicated on each figure. Growth zones are indicated by squares; abbreviations; A = ante- rior, P = posterior. D = dorsal, V = ventral. o^ are estimates of the standard deviation for the observation a at age 1 and A respectively. The third parameter, a, determines nonhnearity of aib), the standard deviation of the observation a. The fourth parameter, /3, controls the extent that the classifica- tion matrix may be dominated by its diagonal en- tries (Richards et al., 1992). Two possible represen- tations of the classification matrix were used, namely a normal and an exponential representation isensu Richards et al., 1992, p. 1803) and the a parameter was constrained to 0 during one fit of each of the normal and exponential representations, allowing four different cases of the model to be fitted for each aging structure. The appropriate model structure, or best fit model, for the classification matrix was then selected by using the Akaike information crite- rion (AIC) where a model with a low AIC value in relation to other models was considered to provide a good fit (Richards et al.. 1992). Initially, data from each aging structure (otoliths, scales, and vertebrae) were analyzed separately and an age assigned to each fish from each structure. The Gillanders et al,: Aging methods for Seriola lalandi 815 Figure 2 Transverse sections of sagittal otoliths of kingfish over a range of sizes viewed with reflected light against a black background: (Al 337 mm FL: (B) 638 mm FL; (C) 850 mm FL; and (D) 1002 mm FL fish. Scale bar is 1.5 mm. classification matrix for the best model (see above) was used to estimate the most probable age of each fish by determining the probability that a fish would be from each true age class given its two assigned ages. Each fish was assigned to the age class with the highest probability (Richards et al., 1992). This estimate of age was used for growth models. Only the first five age classes (0-4) were used in the clas- sification matrix because, although older fish oc- curred, sample sizes were small and some extrapo- lation for missing age classes would be necessary. Because data from older fish are important in esti- mating growth models, fish not used in the classifi- cation matrix were assigned an age by randomly se- lecting one of their two age readings. The data matrix comprising the two age readings for each of the three aging structures was then ex- amined to determine possible effects of the different aging structures and to determine a final age for each fish. Classification matrices for each aging structure were determined following a modification of the methods of Richards et al. (1992; outlined above). Results from running the models for each structure separately showed that the model of best fit was ob- tained by using a normal model with a = 0; there- fore, in its simplest form, the model contained the parameters o^, o^, and terms for the relative bias of each method (rj and rg; r^ - O-r^-r.2). Parameters de- termining the change in bias with age ( y) and non- linearity of bias with age (rj) were also added alone and indexed by aging structure. Different combina- tions of the model parameters were therefore tested to determine which model provided the best fit to the data and to determine relative biases by method. 816 Fishery Bulletin 97(4), 1999 Table 1 Summary of the different cases of Schnute's (1981) size-at-age growth model that was fitted to age estimates from different calcified structures and Francis's ( 1995) mark-recapture analogue of Schnute's growth model that was fitted to the tagging data. Schnute's ( 1981 ) size at age model Francis's (1995) mark-recapture analogue of Schnute's (1981) growth model Case 1: a;tO, b?tO Case 2: a?:0, b^O Case 3: a?iO, b^iO Case 4: sl=^Q, b^tO Meaning of terms Y(t)-- 1 rt~"""'l' Yit) = yjoxp 1-e' 1-e" Y{t)-- y? + (^^yf)^^^ y(<) = ,v,exp log(V2 /.Vi) t-T, y(n = fish size IFL) at age (, Tj and r, = lower and upper ages offish respectively where Tg > fi-.v, and v., = mean sizes at ages r, and Tg, respectively, a and b describe the shape of the cui-ve. AY = -Y,+[Y,''e-''+ca-e-°^)\'' Ay = -y, -I- y;""""^ ' exp[c( 1- e""" )] 1 AY = -Y, +[Y,'' +iX\ - y''^)AtY Ay = -y, + y,(A,/3',)^ Ay = mean growth y, = size at marking y'l and y., = lower and upper sizes of fish, respectively, where y, < v., g, andg., = mean annual growth for fishes of sizes y, and v., respectively. 6 = describes curvature in model, ^1 = yi + gi and A, = Vo + g-,^ a = In An h b y-i - yi if6;tOor ln(y2 /yj) ln(A, /A,) if 6 = 0, c= r- rV^ . iffe;^Oor ln(y)ln(A,)-ln(v,)ln(A„) ._, „ ,, , c = = '—'' ^- 11 o = 0, A? = 1 ln(A,y„)-ln(A., /y,) To determine the timing of zone formation, the edges of the various structures were examined. The growth of the structure, subsequent to the most re- cent zone, was estimated as a percentage (20, 40, 60, and 80%) of the previously completed zone. It was also noted whether the zone was considered to be on the edge of the structure. Only fish with 2-4 gi-owth zones were used. Fish were examined individually by structure in a random order with no knowledge of date of collection. Estimation of growth models from calcified structures Growth models using age estipiates from different calcified structures were derived by using procedures outlined in Schnute ( 1981 ). Schnute's model relates size (FL) to age by several parameters, including two that describe the shape of the curve (a and 6; Table 1 1. These latter parameters combine to describe a range of com- mon growth curves, including the von Bertalanffy (a>0, 6=1), Richards (a>0, 6<0), logistic (o>0, b=-l) and Gompertz (a>0, 6=0) (Table 1; Schnute, 1981). The other parameters in Schnute's growth model were.Vi and Vg, the mean sizes at ages ij and r^ re- spectively, where the value of Tj and r, are specified, but usually chosen to be near the lower and upper ends of the range of ages in the data. In this study, r, and r., were set at 1 and 5 respectively. All growth models were calculated by minimizing sums of squares and using additive errors because variation in size-at-age was similar for all ages of fish (see "Results" section). Initially, a two-parameter model (y, and.v^) was fitted to the data (case 4 in Table 1). Two types of a three-parameter model [parameters were a,y^. and Vp and ^2 (case 3)1 and a four- y.2 (case 2), and 6 Glllanders et al,: Aging methods for Senola lalandt parameter model (a, b, Vj, andvo. case 1) were then fitted to the data (Table 1). To determine whether the addition of extra parameters resulted in a sig- nificantly better fit, significance tests based on the F-distribution were used (Schnute, 1981 ). Where the same number of parameters were present in the models (e.g. comparison of the two models with three parameters ), the model with the lowest residual sums of squares was selected as the best fit. Estimation of rates of growth from tagging data Senola lalandi tagged as part of the NSW Fisheries Gamefish Tagging program (Pepperell, 1985, 1990) were used to estimate growth. A major limitation of these data were that measurement methods were not standard and some measurements appeared spuri- ous. Although most anglers measured total length, some measured fork length only. Where this occurred (17% offish), fork length (mm) was converted to to- tal length with the equation TL (mm)=1.122 x FL + 9. This equation was calculated from fish obtained for aging in which both fork and total lengths were measured (n=570). All fish for which tagging data were available (7?=816) were initially included in analyses even if the data were highly improbable, as with measurements indicating shrinkage between 100 and 350 mm (Fig. 3A). Growth estimates were obtained from the tagging data by using the maximum-likelihood method and the computer program GROTAG (Francis. 1988b). The growth model fitted was Francis's ( 1995) mark- recapture analog of Schnute's (1981) size-at-age model (Table 1). This model provides estimates of^j and ^2' the mean annual growth offish of lengths Vj and ^2 respectively, where Vj and yg ^re chosen to span the range of lengths at tagging. A simple three- parameter model was initially fitted and then addi- tional parameters (growth variability, seasonal growth variation, measurement error, and curvature in the model) were added in a stepwise manner by selecting the parameter that gave the greatest in- crease in log likelihood. At each step, likelihood ra- tio tests were used to determine whether addition of parameters resulted in significantly better fits (Francis, 1988b). Better estimates of the growth pa- rameters can be obtained if measurement error is known (Francis, 1995). There was no way of estimat- ing measurement error without using the current data set; therefore, measurement error was not fixed and this may have compromised estimates of growth. After fitting of the growth model, plots of residuals against length at tagging, time at liberty and ex- pected growth increment were examined for any pos- sible lack of fit of the model. 300 200 100 - i» ^^ffxTi-. i I .rdJJ-n-M-t-L " I ' J3 E o o ooooooooooo oo oooooooo (N,- ^csiro^intof^oo 600 _ Length difference (mm) 400 B 200 - 1 1 Thi-i-j- 1 -1- 1 1 1 1 1 1 II 0 0.5 1 200 r 150 100 50 1.5 2 2.5 3 3.5 4 4.5 Time at liberty (years) irn-i-i-t^-j 200 400 600 800 1,000 1,200 Total length at tagging (50 mm intervals) Figure 3 Distribution of (A) differences in length between tagging and recapture, (B) time at liberty and (C) length-at-tag- ging for kingfish (?!=816). Fish recaptured within 30 days are also indicated in A with shading (n=384). Time at lib- erty (Bl includes a small number of fish recaptured the same day that they were tagged (time at liberty=Ol. Estimation of rates of growth from length- frequency data Length-frequency data were obtained from fish sold at the Sydney Fish Markets between November 1985 and December 1989, prior to the introduction of a size limit of 600 mm TL in February 1990. Data were collected haphazardly amongst months and at locations ranging from 30°S to 37°S. Fork length of fish was measured to the nearest 10 mm and the sampling date and fishing area were recorded. Measurements of ap- proximately 16,000 fish were made, enabling stratifi- cation of samples by month but not by year or area. The von Bertalanffy (VB) model was fitted to the time series of 12 monthly length-frequency distribu- tions by using MULTIFAN software (Fournier et al., 1990, 1991). Likelihood-based methods were used to simultaneously analyze the length-frequency distri- Fishery Bulletin 97(4), 1999 butions sampled at different times to estimate the number of cohorts in the population, the growth pa- rameters (asymptotic length [L^J and growth coeffi- cient [K] of the von Bertalanffy growth equation), the age of the first cohort (assuming that the VB curve passes through the origin) and the proportions at age. The simplest model assumes that mean lengths-at-age lie on the VB gi-owth curve and that the standard deviations of length-at-age are identi- caj for all cohorts. More complex models were also tested that allowed 1) sampling bias for the first co- hort, 2) age-dependent standard deviation in length- at-age, and 3) seasonally oscillating growth to be added as additional parameters in the model. The more complex models incorporated all possible com- binations of these parameters and used likelihood ratio tests to identify the model of best fit. The sig- nificance of improvement of fit within models by add- ing year classes was tested for significance at the 0.10 level ( see Fournier et al., 1990 ), and significance of improvement of fit between models by adding ex- tra parameters was tested at the 0.05 level. The seasonal form of the von Bertalanffy equation for the length-frequency data is 1-p |(j-ll+l/„/12)+/'(/j] l^ja =w,+(;n,v-'"i> ^_ ,,v 1. where f(t„) = —^sin " 2k 12 /j^^ = the mean length of fish of thej'" age class in thea"^ length-frequency data set; Wj = the mean length of the first age class; nij^. = the mean length of the last age class; p = the Brody growth coefficient; t^ = the number of months after the pre- sumed birth month of the fish in the a'*' length-frequency data set; A^ = thenumber of age classes present; and (pj and 02 describe the amplitude and phase of the seasonal component, respectively (Fournier et al., 1990, 1991; Francis and Francis, 1992). Comparisons of rates of growth from calcified structures and from tagging and length-frequency data Rates of gi-owth were estimated from aging struc- tures and from length-frequency (both age-based) and mark-recapture (length-based) data with methods outlined in Francis (1995j. The aging error model of Richards et al. ( 1992) that showed the best fit to the data matrix comprising two age readings from each of the three aging structures was used to determine a single age for each fish. Schnute's growth curve (case 2) was then fitted to the age-size data and this mean age-size relationship was used as a basis with which to compare the estimates of growth from the dif- ferent data sets. The annual growth for the age-size data and length-frequency data was calculated as the mean size at age .v minus the mean size-at-age (.v+1; Francis, 1995 ). Annual growth at corresponding points on the size-based line was then estimated at the size the fish was at age x, as determined from the fit to the mark-recapture data obtained by using GROTAG. Results Structures for aging Seriola lalandi were collected from a total of 572 fish ranging in size from 323 to 1090 mm FL, although not all structures were col- lected from all fish. Although S. lalandi is reported to reach a total length of almost 2000 mm ( 1700 mm FL) and a weight of 60 kg, fish of this size ai-e rare. In New South Wales, few fish over 20 kg ( about 1200 mm FL ) are caught by commercial fishermen and the largest fish recorded in surveys of amateur fishermen has been 1140 mm FL (Steffe et al.-). We were unable to obtain fish larger than 1100 mm FL. All structures showed zones that could be inter- preted as annuli (Fig. 1); however, zones were not interpretable in all fish. Growth zones in whole otoliths were more easily interpretable than those in sectioned otoliths (Fig. 2); the latter showed nu- merous striations that could rarely be interpreted. Zones in sectioned otoliths were, however, clearer in some larger fish (Fig. 2). A large number of scales had to be collected because preliminary results showed that two-thirds offish had at least some re- generated scales. Vertebrae did not always stain well and showed pronounced ridges which reduced read- ability in many fish. Validation of aging methods Analysis of marginal increments showed different patterns among stiaictures (Table 2). Otoliths and scales revealed that one zone was laid down per year, in August-September ( otoliths ) and between Decem- - Steffe, A.. J. Murphy, D. Chapman, B. E. TarHngton, G. N. G. Gordon, and A. Grinberg. 1996. An assessment of the im- pact of offshore recreational fishing in New South Wales on the management of commercial fisheries. F'inal Report to Fisherires Research Development Corporation, PO Box 222. Deakin West ACT 2600, Australia. Gillanders et aL; Aging methods for Seriola lalandi 819 Table 2 Results of analyses of marginal i ncrements for kingfish aged by otoliths, scales, and vertebrae. Each category is | the growth of the structure, subsequent to the most recent zone, as a percentage of the prev iously completec zone. The percentage offish in each category for each two-month | period is shown; t ample sizes are a Iso indicated (in paren- theses beside month). Only fish aged 2-4 were us ed for analyses. For each row, the highest percentage of fish is | shown in bold. Month Category Edge 20 40 60 80 Otoliths Aug-Sep(34) 44 35 15 0 6 Oct-Nov ( 149) 16 36 24 10 13 Dec-Jan (87) 9 7 18 24 41 Feb-Mar (68) 1 13 31 24 31 Scales Aug-Sep (0) Oct-Nov (181) 15 12 21 24 29 Dec-Jan (94) 22 34 21 7 15 Feb-Mar (101) 1 19 28 34 19 Vertebrae Aug-Sep (51) 8 16 16 10 51 Oct-Nov (1731 16 20 24 16 23 Dec-Jan (50) 14 12 28 20 26 Feb-Mar (70) 14 13 37 11 24 ber and January (scales; Table 2). No data were, how- ever, obtained for scales in August-September and sample sizes were small between April and July (n = \- 7 fish per month) for all structures. No clear pattern of marginal zones was observed in vertebrae (Table 2). Precision within and among structures Comparisons of two independent counts of zones in a structure resulted in a relatively low level of agree- ment (50-66'?^^ ). Depending on the structure, between 92% and 96% of readings agreed within one zone (Fig. 4). Differences in counts of zones varied by up to four zones (Fig. 4). Mean coefficients of variation among counts ranged from 7.6% (scales) to 12% (otoliths). Comparison of readings between structures showed a large amount of variation, with differences between structures varying by up to six growth zones (Fig. 5). Agreement between any two methods de- creased with age, but otoliths and vertebrae had the greatest concordance in fish aged 4 and over. Fish were never assigned an age of 0 when aged with scales; fish assigned 0 or 1 with other structures were assigned an age of 1 with scales. Readings between structures agreed within one zone between 88% (be- 300 200 100 400 i 300 o I 200 E z 100 0 300 200- - A (n=475) 1 1 1 1 1 1 1 III 1 Scales „ - (n=481) JB 1 1 . — I— 1 1 1 — 1 — . 1 1 Vertebrae c 1 1 1 , 1 , — 1 — , 1 1 -4-3-2-101234 Difference (number of zones) Figure 4 Differences between repeated counts of zones for three structures used to age kingfish. Each comparison repre- sents independent counts from a single reader tween vertebrae and otoliths) and 91% (between scales and otoliths) of the time. Estimates of aging error Analysis of aging precision with the methods of Richards et al. (1992) showed that the estimates of assigned age were more precise for young fish (Fig. 6A). For example, with otoliths, 98% offish with an estimated most probable age of 0 were likely to be aged as 0, whereas only 74% offish with a most probable age of 4 were likely to be aged as 4. This precision can be compared with those obtained from other structures (e.g. scales and vertebrae; Fig. 6A). Fish aged with scales had a higher probability of being consistently assigned ages 1 and 2. Fish aged with vertebrae or otoliths, however, had a higher probability of being consistently assigned age 4, whereas there was little difference between the three structures in assigning fish to age 3 (Fig. 6A). 820 Fishery Bulletin 97(4), 1999 A more complex aging error model, with para- meters that accounted for effects due to aging struc- tures showed that fish in their first year (i.e. 0 growth zones) were likely to be overestimated with scales and vertebrae than with otoliths. Otoliths underes- timated the age of fish in their first year, although this is due to the relative nature of the model be- cause vertebrae and scales overestimated the age of these fish (Fig. 6B). Scales showed less bias in rela- tion to the other structures for fish aged 1^ (Fig. 6B). Estimates of rates of growth Patterns in growth of kingfish from NSW were ob- tained from Schnute's (1981) growth model fitted to the estimated most probable ages of fish calculated 200 Otoliths & scales ^^ ; n=425 10 - 150 0) o to 1 iX' 1 ..>■' 100 - 1^ 1 1 1 •;* 1 1 3 3 t"l 1 "' 4 7 163d'4 1 1 50 1 2 - 1 39^.38' 10 ? 43^v 54 10 2 sja'e 1 C^IIZ) 0 -6-4-2024 " 2 4 6 8 10 12 Otolith age | 200 Otoliths & vertebrae : n=353 , ^ 10 1 ,.. 1 150 §> f-' Z ^ 2 1 /" o 2 1 1 ,i 1 100 1 6 ' ,'2 E ^ 6 4,'l 1 ^ 50 , , , ,r _i a 4 2 Tl 1 1 a 19,^4 1 1 - 2 49^ 15 4 6 10,^34 4 6,« 11 1 ° -6 -4 -2 0 2 4 "' 2 4 6 8 10 12 Otolith age 200 Scales & vertebrae ^ ^ I n=356 , r-1 0)10 ^ , '-'■ 150 r ' ' ',.;■■'■■" 4 1 1 , ■■ 100 1 6 2 3-1 ■c 4 3 ,1 1 m A ~1 ^ 10 ^2,fi 4 2 50 ] 2 -n , . - 1 69,*l 11 r 6 ,^8 24 6 1 -^19 0 -6-4-2024 " 2 4 6 8 10 12 1 Difference (no. zones) Scale age Figure 5 Counts of zones from kingfish otoliths, .scales, and vertebrae com- pared with other aging structures. Histograms show the differ- ence between structures, and the scatterplots enable compari- sons between structures by age. In each scatterplot, the line of slope 1 (dotted line) indicates 100% agreement. The line of best fit (dashed line) is indicated with a dashed line. The sample size for each point on the scatterplots is also shown. from the different aging structures (Fig. 7). Size-age data were best fitted by three-parameter models for otoliths, vertebrae (both case 2, see Table 1) and scales (case 3). Using the best fitting model for each structure, we estimated that the average lengths (± standard error) offish with one growth zone were 499 (±5), 418 (±9), and 485 (+7) mm FL for otoliths, scales, and vertebrae respectively. At age 5, fish were 807 (±8 ), 823 (±8), and 788 (±6) mm FL for otoliths, scales, and vertebrae respectively. The maximum age offish was 9 yr for otoliths and scales, compared to 1 1 yr for verte- brae (Fig. 7). Vertebrae tended to produce higher age estimates than did otoliths and scales, with 5% offish showing greater than five growth zones in vertebrae compared with only 27( for scales and otoliths. Using a single age for each fish, we estimated an age-length key for fish sampled during the current study (Table 3). Fish in all length classes were found in more than one age class (Table 3). The different aging structures showed signifi- cant differences in mean size-at-age for some age classes. Scales showed significantly differ- ent mean size-at-age to otoliths and vertebrae for fish with one growth zone (^-tests, P<0.05). Otoliths and vertebrae gave similar estimates of mean size-at-age for fish with less than eight growth zones (<-tests, P<0.05). With the excep- tion of fish with one growth zone, scales gave similar estimates of mean size-at-age to otoliths for all other age classes (t-tests, P<0.05). Al- though differences between scales and vertebrae were statistically significant for some age classes (fish with 4—6 growth zones), these differences were not likely to be biologically important. Kingfish that were measured at both tagging and recapture were at large for between 0 days (i.e. recaptured the same day that they were tagged) and 5 years (Fig. 3B). Growth of recap- tured fish ranged from a decrease of 350 mm to an increase of 800 mm (Fig. 3A). The frequency distribution of fish at large <30 days showed that fishermen were just as likely to underesti- mate the length offish as they were to overes- timate the length of fish (Fig. 3A). The mean difference in size of fish between tagging and recapture for fish recaptured within 30 days was 11.6 mm (±0.4, SE), suggesting that there was little bias in measurements between tagging and recapture. Size offish at tagging ranged from 220 to 1200 mm, although the majority of fish were between 400 and 600 mm TL (Fig. 3C). The best fit to the complete tag-recapture data set (model 1 in Table 4) showed a high pro- portion of outliers (P=0.04). Twenty-two fish (or 2.7^^) that had absolute standardized residu- Gillanders et al : Aging methods for Seriola lalandi 821 als greater than 3 in the model 1 fit were removed from the data set. Initially, a simple three-param- eter model was fitted (model 2 in Table 4), which indicated growth rates of 255 mm and 176 mm for 400-mm and 600-mm fish, respectively. The first additional parameter selected was that describ- ing the shape of the growth curve (model 3 in Table 4). Model 3 showed annual growth rates of 270 and 140 mm. There was a significant improvement in fit when a term describing growth variability was added (model 4 in Table 4). Additional parameters (e.g. seasonal growth terms) did not result in sig- nificant improvements of fit. The best fitting model was therefore case-1 model in Table 1. Plots of standardized residuals against length at tagging, time at liberty, and expected growth increment showed no pattern (correlations were 0.022, 0.003, and 0.005, respectively) suggesting that the model was appropriate. The best fitting model for the kingfish length-fi-e- quency data set identified five cohorts aged from 1.73 to 5.73 years (Table 5; Fig. 8). The standard devia- tion of the predicted length estimates for each age class increased with age. Parameters for first length bias and seasonal growth were also included in the best fitting model (Table 5; Fig. 9). Estimates of the mean length-at-age ( and standard deviation ) for the length-frequency data set are given in Table 5. The mean lengths and proportions of the modes predicted by MULTIFAN generally fitted the observed data I Fig. 8). There were large numbers of small fish (e.g. 1 yr) in the catch during the summer months (e.g. November and December), whereas 2-yi- fish dominated the catch at other times. Small numbers of large fish (e.g. gi-eater than 5 yr) were found throughout the year (Fig. 8). The seasonal form of the von Bertalanffy growth equation showed that the projected value of L,, was 1252 mm FL and the rate of change in growth increment (K) was 0.189 (Fig. 9). Comparison of annual growth between age- based (age-length and length-frequency) and length-based (mark-recapture) data, although not strictly comparable, showed a decrease in growth with age and size offish (Fig. 10). Estimates of annual growth were similar among the three methods for 2-4 year fish (=550-750 mm SL) but varied by =50 mm for 1-yr-old fish. Discussion All structures showed patterns of growth that were, to varying degrees, quantifiable. Delineation of each zone was, however, sometimes difficult, as has been q > ■o P 1.0,- 08 0.6 0.4 0.2 0.0 1.0 — Vertebrae ■ ■ Scales — Otoliths _L _L 0.5 - -0.5 -1.0 12 3 4 Estimated most probable age (no. zones) Vertebrae D Scales Otoliths r _L _L 0 12 3 4 Estimated most probable age (no. zones) Figure 6 (A) Aging error versus estimated most probable age for king- fisli aged with otoliths, scales, and vertebrae. Estimated most probable age was calculated separately for each structure by using the methods of Richards et al. ( 1992), see text for fur- ther details. Errors around the first and last ages are indi- cated by the vertical lines. (B) Relative bias versus estimated most probable age for kingfish aged by otoliths, scales, and ver- tebrae. Relative bias was calculated from a model that incorpo- rated two age readings from each of the three structures. found in other studies (e.g. Brennan and Cailliet, 1989; Manooch and Potts, 1997), and the clarity of zones varied among individuals for all structures. Because pelagic fishes including Seriola spp. are known to be difficult to age, these results were not surprising. To accurately reflect the age of a fish, the zones must be formed on a regular and determinable time scale. 822 Fishery Bulletin 97(4), 1999 .^ Small sample sizes during the winter months (from April to July), due in part to small numbers offish be- ing taken in the fishery, made analysis of marginal in- crements problematic, especially for scales because samples were also not obtained in August-September. With the exceptions of otoliths and possibly scales, our data were not sufficient to confirm that only one zone otoliths fl2=0.67 SD=72 36 a=0 1322(0.0246) y1=499(5) y2=807 (8) J I I I 1.200 1.000 800 600 400 200 0 8 9 10 11 12 Scales H2=0.73 SO=66 35 6=2 408(0 203) y1=418 (9) y2=823 (8) I I I I I 0 1 8 9 10 11 12 Vertebrae R2=0 70 SD=64 50 a=0 1844 (0 0243) y1=485(7) y2=788 (6) 10 11 12 Growth! curves 4 5 6 7 8 9 Age (no, zones) 10 11 12 Figure 7 Relationship between fork length and age for the different struc- tures used to age kingfish (A-C) and (D) comparison of the three growth curves. The growth curves were calculated by using Schnute's growth model. /?-, overall standard deviation (SD), and the parameters (standard error) describing the growth model are shown. _V[ is size at age 1, y.^ is size at age .5, a and b are parameters that describe the shape of the curve. The for- mulae for the growth curves are found in Table 1. was formed per year. No single category of marginal growth ever had more than 50% of fish for any struc- ture. A previous study (Mitani and Sato, 1959) also found that fish (Seriola quinqueradiata) collected in any one time period showed a wide range of marginal growth conditions. Growth zones on the edge of the otoliths of S. dumerili were also found over five months of the year (Manooch and Potts, 1997). The wide range of marginal growth patterns in our study may in part be due to grouping of samples into two-month intervals or because fish from three age classes (2—4 growth zones) were used in analyses. Fish from different age classes have previously been shown to lay down zones at slightly different times of year (e.g. Francis et al., 1992). Analysis of marginal increments often provides only partial validation of age estimates because older fish with slower growth often do not show seasonality in formation of zones. A method should not be considered accurate until all re- ported ages are validated (Beamish and McFar- lane, 1983). Validation of age estimates in older fish will require a mark-recapture study, but fur- ther work on validation in younger fish may use a variety of approaches (e.g. length-frequency analyses if cohorts are easily recognized, mark- recapture etc). In the present study, few tagged fish were at large more than a year (Fig. 3B), which suggests that many fish would have to be tagged in a mark-recapture study to recover suf- ficient fish to make this approach feasible. There have been few validation studies on Seriola spp. and all have involved analysis of marginal in- crements. Mitani and Sato (1959) have suggested that one zone is laid down each winter in opercular bones ofS. quinqueradiata and Baxter ( 1960) foimd that zones are formed between November and January in scales of S. lalandi (formerly dorsal is). In otoliths of S. lalandi opaque zones appear to be laid down in August or September Timing of zone formation in scales and vertebrae was more vari- able and may differ between structures because pro- cesses involved in deposition vary among bone, scales, and otoliths (Simkiss, 1974). Precision of aging estimates Exact agreement among age estimates for each structure was generally poor; agreement within one zone was reasonable for all structures. The use of percent agreement (i.e. percent offish aged alike between sets of multiple readings) has been criticized because it fails to take into account the range of year classes offish and therefore can be used only for age-specific comparisons (PCimura Gillanders et al,; Aging methods for Ser/o/a lalandi 823 Table 3 Kingfish age -length (FL, mm) distribution for fish collected from New South Wales, Australia, between August 1995 an d July 1996. Shown are the nt mber offish and the percentage of the length c ass (in parentheses) Ages were estimated from the aging error model that showed the best fit to the data matrix comprismg two age reading 5 from each of the three aging structures where fish had less than five growth zones or by randomly selecting one of the six age readings where fish had five or more growth zones. Length class Age (yr) 0 1 2 3 4 5 6 7 8 9 Total 300-350 4(66.7) 1(16.7) 1(16.7) 6 351-400 11(47.8) 7(30.4) 5(21.7) 23 401-450 4(40.0) 4(40.0) 2 (20.0) 10 451-500 11(34.4) 19(59.4) 2(6.3) 32 501-550 . 11(10.6) 85(81.7) 8(7.7) 104 551-600 1(1.1) 67(71.3) 26(27.7) 94 601-650 54 (50.9) 51(48.1) 1 (0.9) 106 651-700 14(19.2) 52(71.2) 6(8.2) 1(1.4) 73 701-750 5(13.2) 16(42.1) 14(36.8) 2 (5.3) 1(2.6) 38 751-800 1 (3.1) 9(28.1) 16(50.0) 4(12.5) 1(3.1) 1(3.1) 32 801-850 3(21.4) 4(28.6) 4(28.6) 3(21.4) 14 851-900 2 (62.5) 5(62.5) 1(12.5) 8 901-950 1 (18.2) 2(18.2) 4(36.4) 3(27.3) 1(9.1) 11 951-1000 1(33.3) 1(33.3) 1(33.3) 3 1001-1050 1(33.3) 1(33.3) 1(33.3) 3 1051-1100 1(33.3) 2(66.7) 3 N 19 35 253 167 44 19 12 4 3 4 560 Table 4 Log likelihood function values, growth parameter estimates, and standard errors (model 4 only) for kingfish fSeriola lalandi) tagging data. * indicates that parameters were held fixed. Standard errors of parameter estimates were estimated from simu- lated data (n=100 simulations). Growth rates are shown for 400 mm (g4oo) and 600 mm (g6oo> TL fish. The best model fit (model 4) is the case-1 model from Table 1. Parameter Model 1 Model 2 Model 3 Model 4 SE Log-likelihood 2890.67 2797.80 2735.35 2706.21 Mean growth rate g40o20 kg). Comparison of age- and length-based data Although not strictly comparable (see Francis, 1988a, 1995), estimates of growth from length-frequency, 826 Fishery Bulletin 97(4), 1999 age-length (both age-based data), and tagging data (length-based) showed agreement for fish aged 2-4 years but varied for fish with one growth zone. Dif- ferences in rates of growth were greater for younger fish than for older fish and may have been caused by inaccuracies in aging, influence of tagging on growth (e.g. McFarlane and Beamish, 1990), within- or between-year differences, and variations in year-class strength. Although estimates of growth from tagging data provide some indication that age-length data 1.000 800 600 400 200 0 _ ^ ^—^^^ -""^ fo = -0,735 y^ /<=0,189 1 1 1 1 /-,, = 1 .252 1 1 1 3 4 Age (years) Figure 9 MULTIFAN growth curve for the length-frequency data. A model incorporating seasonal growth is shown. Values of ^i, A', and L , (mm) are shown. 200 r- , 1 150 - X. Size frequency Annual growth ( o o '\\ \^ \ ~ "~^-?r ^^^^-^..^ 50 1 1 1 1 1 1 1 1 12 3 4 5 6 7 8 Age (no zones) ! 1 1 1 1 1 ! 400 500 600 700 800 900 100 0 Fork length (mm) Figure 10 Comparison of rates of annual growth for kingfish estimated from lA) aging structures, (B) length-frequency (both age-based) and (Ci mark-recapture (length-based) data. The mean length at any age is aligned with the corre- sponding point on the age axis; this was calculated from age-size data ob- tained from the aging structures. may be reasonable, they should not be used as a means of validation (see Francis, 1988a). Tagging data also suggest that S. lalandi in New South Wales has greater annual growth than the same species in New Zealand and the United States (Baxter, 1960; Holdsworth^). For S. lalandi in New South Wales, annual growth rates of 144 mm were found for 500-mm-FL fish, compared with 93 mm (New Zealand) and a range of 34-109 mm (US) for similar size fish. In New Zealand, annual growth of 44 mm was found for 1000-mm-FL fish, which is within the range found in the United States ( 19- 70 mm). Few large fish were tagged in New South Wales, preventing estimates of annual growth at this larger size. Practicalities of aging kingfish With the exception of dorsal spines, which did not appear useful for aging kingfish, only scales can be easily collected without altering the market value of the fish because Seriola lalandi are pref- erentially sold whole in NSW. Scales showed simi- lar estimates of mean size-at-age to otoliths for all age classes except fish with one growth zone, but their use in aging kingfish can only be recom- mended with caution until the position of the first gi'owth zone is better understood. The usefulness of otoliths and vertebrae may be limited by the cost offish, but, if possible, col- lections of at least one of these struc- tures should be made. The similarity of age estimates among structures sug- gests that, if validations are possible, kingfish may be aged reliably. Acknowledgments This research was supported by the Fisheries Research Development Corpo- ration and NSW Fisheries. We thank the processors at the Sydney Fish Mar- kets, commercial fishermen, NSW Fish- eries staff and Martin Tucker for supply- ing fish. We also thank Norm Lenehan for help with processing vertebrae, and Duncan Worthington and Malcolm Haddon for statistical advice and nu- merous discussions regarding aging of fish. We are particularly grateful to Holdsworth, J. C. 1997. Ministry of Fisher- ies, 17 Keyte Street, Kensington, private Bag 901.3, Whangarei. New Zealand. Unpubl. data. Gillanders et al.: Aging methods for Seriola lalandi 827 Duncan Worthington for his help and advice in run- ning MATLAB for the aging error section of this pa- per and for his comments on a draft of the manu- script. The manuscript was greatly improved by the comments of three anonvmous reviewers. Literature cited Bagenal, T. B., and F. W. Tesch. 1978. Age and growth. Chap. 5 in T. B. Bagenal (ed. ), Meth- ods for assessment offish production in fresh waters, p. 101- 136. Blackwell Scientific Pubhcations, Oxford, 365 p. Baxter, J. L. 1960. A study oftheyellowtailSeno/ac?orsa/(s (Gill). State of California Department of Fish and Game, Fish Bull. 110:96. Beamish, R. J. 1979. Differences in the age of Pacific hake (Merluccius productus) using whole otoliths and sections of otoliths. J. Fish. Res. Board Can. 36:141-151. Beamish, R. J., and G. A. McFarlane. 1983. The forgotten requirement for age validation in fish- eries biology. Trans. Am. Fish. Soc. 112:735-743. 1987. Current trends in age determination methodology. In R. C. Summerfelt and G. E. Hall (ed.). Age and growth of fish. p. 15-42. Iowa State Univ. Press, Ames, lA, 544 p. Berry, F. H., D. W. Lee, and A. R. Bertolino. 1977. Age estimates in Atlantic bluefin tuna — an objective examination and an intuitive analysis of rhythmic mark- ings on vertebrae and in otoliths. Int. Comm. Conserv. Atl. Tunas, Collect. Vol. Sci. Pap., Madrid 6:305-317. Brennan, J. S., and G. M. Cailliet. 1989. Comparative age-determination techniques for white sturgeon in California. Trans. Am. Fish. Soc. 118:296-310. Campana, S. E. 1984. Comparison of age determination methods for starry fiounder Trans. Am. Fish. Soc. 113:365-369. Foumier, D. A., J. R. Sibert, J. Majkowski, and J. Hampton. 1990. MULTIFAN a likelihood-based method for estimat- ing growth parameters and age composition from multiple length frequency data sets illustrated using data for south- ern bluefin tuna iThunnus maccoyii). Can. J. Fish. Aquat. Sci. 47:301-317. Foumier, D. A., J. R. Sibert, and M. Terceiro. 1991. Analysis of length frequency samples with relative abundance data for the Gulf of Maine northern shrimp (Pandalus borealis) by the MULTIFAN method. Can. J. Fish. Aquat. Sci. 48:591-598. Francis, M. P., and R. L C. C. Francis. 1992. Growth rate estimates for New Zealand rig (Mustelus lenticulatus). Aust. J. Mar. Freshwater Res. 43:1157- 1176. Francis, R. L C. C. 1988a. Are growth parameters estimated from tagging and age-length data comparable? Can. J. Fish. Aquat. Sci. 45:936-942. 1988b. Maximum likelihood estimation of growth and growth variability from tagging data. N.Z. J. Mar Fresh- water Res. 22:42-51. 1995. An alternative mark-recapture analogue of Schnute's growth model. Fish. Res. 23:95-111. Francis, R. I. C. C, L. J. Paul, and K. P. Mulligan. 1992. Ageing of adult snapper iPagrus auratus) from otolith annual ring counts: validation by tagging and oxytetracycline injection. Aust. J. Mar Freshwater Res. 43:1069-1089. Hoenig, J. M., M. J. Morgan, and C. A. Brown. 1995. Analysing differences between two age determina- tion methods by tests of symmetry. Can. J. Fish. Aquat. Sci. 52:364-368. Kimura, D. K., and J. J. Lyons. 1991. Between-reader bias and variability in the age-de- termination process. Fish. Bull. 89:53-60. Manooch, C. S. Ill, and J. C. Potts. 1997. Age, growth and mortality of greater amberjack from the southeastern United States. Fish. Res. 30:229-240. McFarlane, G. A., and R. J. Beamish. 1990. Effect of an external tag on growth of sablefish iAnoplopoma fimbria I, and consequences to mortality and age at maturity. Can. J. Fish. Aquat. Sci. 47:1551-1557. Mitani, F. 1955. Studies on the fishing conditions of some useful fishes in the western region of Wakasa Bay — 11. Relation between the scale of the yellowtail, Seriola quinqueradiata T. & S., and fork length and age. Bull. Jpn. Soc. Scient. Fish. 21:463-466. 1958. Studies on the growth and age of the yellow-tail, Seriola quinqueradiata T. & S., found in Japan and the adjacent region — 1. A fundamental consideration on age estimation by vertebrae. Bull. Jpn. Soc. Scient. Fish. 24:626-631. Mitani, F., and T. Sato. 1959. Studies on the growth and age of the yellowtail, Seriola quinqueradiata T. & S., found in .Japan and the adjacent re- gion— 11. Estimation of age and growth from the opercular bone. Bull. Jpn. Soc. Scient. Fish. 24:803-808. Munekiyo, M., M. Sinoda, and O. Sugimura. 1982. A possibility offish age estimation by means of a rep- lica of the vertebral centrum. Bull. Jpn. Soc. Scient. Fish. 48:1371-1374. Murayama, T. 1992. Growth of the yellowtail Seriola quinqueradiata in the Japan Sea and coast in recent years. Nippon Suisan Gakkaishi 58:601-609. Pepperell, J. G. 1985. Cooperative game fish tagging in the Indo-Pacific region. In R. H. Stroud (ed.). World angling resources and challenges, p. 241-252. International Game Fish Asso- ciation, Fort Lauderdale, Florida. 1990. Australian cooperative game-fish tagging program, 1973-1987: status and evaluation of tags. Am. Fish. Soc. Symp. 7:765-774. Richards, L. J., J. T. Schnute, A. R. Kronlund, and R. J. Beamish. 1992. Statistical models for the analysis of ageing error. Can. J. Fish. Aquat. Sci. 49:1801-1815. Schnute, J. 1981. A versatile growth model with statistically stable parameters. Can. J. Fish. Aquat. Sci. 38:1128-1140. Simkiss, K. 1974. Calcium metabolism of fish in relation to ageing. In T. B. Bagenal (ed.), The ageing of fish, p. 1-12. Unwin Brothers Ltd, London, 234 p. 828 Abstract.— Fishing pressure on ver- milion snapper, Rhomboplites auro- rubens, in the Gulf of Mexico has in- creased since the mid-1970s, and popu- lations are thought to be overfished. We sampled 858 vermilion snapper (192- 585 mm TL) from the eastern Gulf of Mexico during 1995 and 1996 to assess their age structure, growth, mortality, spawning season, size and age at ma- turity, and batch fecundity. The lengths of males and females in our samples were not significantly different, and the overall sex ratio was not significantly different from 1:1. Marginal-increment analysis indicated that one opaque zone is formed on vermilion snapper otoliths during the late spring to early summer of each year Ages ranged from 1 to 13 years, and von Bertalanffy growth mod- els for males and females were not sig- nificantly different. Von Bertalanffy growth model parameters were L„ = 298 mm TL, A'=0.25/yr, and /o=-3.9 years for all aged fish. Growth rates in our study were lower than those in pre- vious studies of Gulf of Mexico vermil- ion snapper, perhaps the result of re- cent changes in fishing selectivity. Pooled estimates of total instantaneous mortality were 0.480/yr based on rec- reational landings data and 0.489/yr based on commercial landings data. Most females and all males examined were mature. At 200 mm TL, 90% of the females we examined were mature. Vermilion snapper are summer spawn- ers, and ripe females were caught from May to September. Batch fecundity ranged from 5535 to 22,81 1 oocytes and was positively correlated with fish weight (batch fecundity=317 x (whole weight) - 3.1624 x 10^ r-' = 0.55). Age, growth, mortality, and reproduction of vermilion snapper, Rhomboplites aurorubens, from the eastern Gulf of Mexico Peter B. Hood Florida Marine Research Institute, Florida Department of Environmental Protection 100 Eighth Avenue S,E, St. Petersburg, Florida 33701-5095 Present address: Gulf of Mexico Fishery Management Council 3018 U.S Highway 301 North, Suite 1000 Tampa, Flonda 33619-2266 E-mail address peter hoodig^gulfcouncil org Andrea K. Johnson North Carolina State University, Department of Zoology Box 7617 Raleigh, North Carolina 27695-7617 Manuscript accepted 16 November 1998. Fi.sh. Bull. 97:828-841 (1999). The vermilion snapper, Rhombo- plites aurorubens, is a small, sub- tropical lutjanid that occurs from North Carolina to Rio de Janeiro but is most abundant off the south- eastern United States and in the Gulf of Campeche (Vergara, 1978). In the Gulf of Mexico (GOM), ver- milion snapper are usually found near hard bottom areas off the west- central Florida coast, the Florida Middle Ground, and the Texas Flower Gardens (Smith et al., 1975: Smith, 1976; Nelson, 1988), Faunal surveys in the South Atlantic Bight (SAB) have indicated that vermil- ion snapper are most common over inshore live-bottom habitats and over shelf-edge, rocky-rubble, and rock-outcrop habitats (Grimes et al., 1977, 1982; Barans and Henry, 1984; Chester et al., 1984; Sedberry and Van Dolah, 1984). Vermilion snapper are an impor- tant component of the reef fish fish- ery on the west coast of Florida, In 1995 and 1996, total commercial landings of vermilion snapper on Florida's west coast was 1,6 million pounds and had an estimated dock- side value of $4.5 million (Marine Fisheries Information System' ). Over the same period, 1.8 million vermil- ion snapper were landed by anglers in Flonda (MRFSS2), Charter boats or headboats accounted for most ( 90% ) of the angler-caught fish ( Good- year and Schirripa'^ ; Schirripa'*). The Gulf of Mexico stock of ver- milion snapper is showing signs of being overfished, Schirripa'* re- ported in an analysis of GOM ver- milion snapper landings data that 1) overall landings were declining, 2) the fishery was consolidating to the most productive fishing areas, 3) the mean size offish in the com- mercial catch was decreasing, 4) the catch per unit of effort was decreas- ' Marine Fisheries Information System. 1997. Florida Department of Environ- mental Protection, 100 8th Avenue S.E., St. Petersburg, FL. Unpubl. data. ^ MRFSS (Marine Recreational Fishery Sta- tistics Survey). 1997. Fisheries Statis- tics Division, National Marine Fisheries Service, Department of Commerce, Silver Springs, MD. Unpubl. data. ' Goodyear, C. P., and M. J. Schirripa. 1991. A biological profile for vermilion snapper with a description of the fishery in the Gulf of Mexico. Miami Laboratory, Southeast Fisheries Center, National Ma- rine Fisheries Service, NOAA, Miami, FL. Contribution rep. MIA-90/91-78. ■* Schirripa, M. J. 1996. Statusof the ver- milion snapper fishery of the Gulf of Mexico: assessment 3.0. Miami Laboratory, Southeast Fisheries Science Center, Natl. Mar Fish. Serv., NOAA, Miami, FL. Con- tribution rep. MIA-9.5/96-61. Hood and Johnson: Life history of Rhombophtes aurorubens 829 ing, and 5) the estimated number of age- 1 fish in the population was declining. Goodyear and Schimpa'^ also noted that large differences in estimates of length-at- age and estimates of fecundity made it difficult to as- sess confidently conditions of the stock by using stan- dard assessment models. Existing age and growth data for this species in the GOM are inadequate and dated. Zastrow ( 1984) and Nelson (1988) used scales to age GOM vermil- ion snapper collected in the early 1980s. The maxi- mum age reported was 7 years. However, using scales to determine age has proven to be problematic. Stud- ies in the SAB have shown that as age increases, the readability of the scales decreases (Grimes, 1978; Collins and Pinckney, 1988). Barber (1989) used whole otoliths and scales to age GOM fish and was able to count up to 18 rings in scales and 26 rings in whole otoliths. Although he did not directly compare ages from both structures, his scale- and otolith- based estimates of length at age and growth were very different. In addition, his attempt to validate whole-otolith-based ages was unsuccessful. Growth of vermilion snapper may be affected by changes in fishing. Zhao et al. ( 1997) found that mean and predicted sizes-at-age of SAB fish had declined between 1979 and 1993 and associated this change with size-selective fishing. They suggested that fish- ing is removing faster-growing fish from the popula- tion and may have genetic or physiological conse- quences in the life history of this species. Information on the reproductive biology of vermil- ion snapper in the GOM is limited. Sex ratio appears to be dependent on location. Sex ratios from the GOM and Puerto Rico are approximately 1:1 (Boardman and Weiler, 1979; Zastrow, 1984; Collins''') although Nelson (1988) reported that males outnumbered fe- males 1.2:1. In the SAB, females consistently out- numbered males, and sex ratios ranged from 1.6:1 to 1.7:1 (Grimes and Huntsman, 1980; Collins and Pinckney, 1988; Cuellar et al., 1996; Zhao and McGovern, 1997). Nelson (1988) examined spawn- ing period, fecundity, and sex ratios offish caught in the western GOM. His results were similar to find- ings reported for the SAB (Grimes and Huntsman, 1980; Cuellar et al., 1996). In both regions, spawn- ing occurs during the summer and early fall. Nelson (1988) estimated that vermilion snapper batch fe- cundities in the GOM range from 61,600 to 392,000 eggs. Fecundity has been found to have a positive relationship with fish size (Grimes and Huntsman, 1980; Nelson, 1988; Cuellar et al., 1996). Vermilion ■' Collins, A. 1997. Southeast Fisheries Science Center, Natl. Mar. Fish. Serv., 3.500 Delwood Beach Road., Panama City, FL 32407. Personal commun. snapper are thought to spawn in aggregations. Boardman and Weiler ( 1979) and Grimes and Hunts- man ( 1980) found large numbers offish in the same reproductive state in single collections. Basic life-history information is needed to prop- erly assess vermilion snapper stocks in the GOM. Accurate ages are needed to develop age-length keys, develop growth models, and estimate total mortal- ity. In addition, the annual periodicity of ring depo- sition in otoliths has not been validated in the GOM. With the increasing reliance on estimates of spawn- ing-potential ratios to describe a stock's condition, information on maturation schedules, sex ratios, and size-specific fecundities are also needed. The purpose of this study was to accurately age eastern GOM ver- milion snapper to develop age-length keys, develop growth models, construct catch curves for deriving estimates of total mortality, and to describe the re- productive biology of this species. Methods Eastern GOM vermilion snapper were sampled from October 1995 to September 1996. Samples were ob- tained from headboat fishermen, commercial catches, and a Florida Department of Environmental Protec- tion trawl survey. Total length (TL). fork length (FL), and standard length (SL) were measured to the near- est millimeter. Whole weight or gutted weight (or both) were measured to the nearest gram. The rela- tionships between lengths and logjo-transformed to- tal lengths and weights were determined by least- squares regression (SAS Institute, Inc., 1985). Male and female regression lines of logjg-transformed to- tal lengths and weights were compared by using analysis of covariance (Snedecor and Cochran, 1971). Thin sections of sagittae (hereafter referred to gen- erally as otoliths) were used to determine the ages offish. Otoliths were removed and stored dry in cul- ture wells. The left otolith was serially sectioned across its anterior-posterior midpoint at 0.5-mm in- tervals by making a transverse cut with an Isomet diamond saw. Mounted sections were placed on a black field, illuminated with reflected light, and ex- amined with a binocular dissecting microscope. The magnified images of otolith sections were transmit- ted by means of video camera to a video monitor and were analyzed with a computer-driven data-acquisition software package (Optimas Corp., 1996). The number of opaque zones and the radial measurements from the core to the last opaque band and to the edge of the otolith (otolith radius. Fig. 1) were recorded. Marginal increments were measured as the distance between the last opaque band and the edge of the otolith. 830 Fishery Bulletin 97(4), 1999 To determine the precision of vermilion snapper otolith opaque zone counts, otoliths collected during the first four months of sampling (n=200) were read independently by two investigators. After the first reading, readers examined the sections together and compared counts to form a consensus about what constituted an opaque zone. The two readers then re-examined the sectioned otoliths independently, and counts were compared again. Because agreement between readers was 100%, one reader read the re- maining otoliths to determine age. Each of the re- maining otolith sections was read three times, and opaque zone counts were accepted for ages only if at least two of the three separate readings were the same. To validate annulus periodicity, marginal in- crements and their medians were plotted by month for each age and compared for consistent temporal patterns. Age in years was estimated as the number of opaque zones. Therefore, length at age included any growth that occurred after the last opaque ring was formed. Mean length at age was calculated for males, females, and all aged fish. Age and observed length data were fitted to a von Bertalanffy growth model by using a nonlinear regression (SAS Institute, Inc., 1985). Growth curves were fitted separately for males, females, and all aged fish. The estimated pa- rameters for male and female curves were compared by using likelihood-ratio tests (Kimura, 1980; Cerrato, 1990). We calculated an adjusted r-^ for the re- sulting curves by using methods described by Helland (1987). We used age-frequency data from this study to estimate mortality rates from the commercial and rec- reational length data. Because the numbers of vermilion snapper mea- sured in fishery sampling programs were low, we pooled the most recent years where data were available. Length data from commercial land- ings were obtained from the TIPS*^ from 1992 to 1994, and in 1996. Length data fi-om recreational land- ings were obtained from the MRFSS- and pooled for years 1990-96. Instan- taneous mortality and survivorship were estimated by the Chapman- Robson method (Youngs and Rob- son, 1978). Age at full recruitment was estimated from the catch curve as the next oldest age from the age with the greatest catch. Reproductive analyses were based on gonad weights and a histological examination of gonadal tissue. Whole gonads were weighed to the near- est 0.1 g, fixed in 10% buffered for- malin for approximately one week, rinsed in water, and then trans- ferred to 70% ethanol. A sample was taken from the middle of the pre- Figure 1 Sectioned saggitae from (A) a 2-year-old (204-mm-TL) and (B) a 10-year-old ( 184-m-TL) vermilion snapper. Both fish were caught in the eastern Gulf of Mexico in July 1996. •^ TIPS (Trip Interview Program). 1997. Florida Department of Environmental Pro- tection, 100 8th Avenue, S. burg, FL. Unpubl data. E., St. Peters- Hood and Johnson: Life history of Rhomboplites aurorubens 831 Table 1 Gonad development classes for vermilion snapper Class Ovary Testes Immature Only primary growth oocytes present. Spermatogonia and spermatocytes in the central lobules; no tailed sperm in the lumen of the lobule. Resting Primary growth oocytes and atretic bodies present. Mostly spermatogonia and spermatocytes in the central lobules. Free spermatozoa in the lumen of the lobule, and brown bodies (drier, 1987) may be present. Early developing Primary growth and cortical alveoli (yolk vesicle) oocytes present. Atretic bodies may be present. Developing Primary growth, cortical alveoli, and vitellogenic oocytes present. Mostly spermatocytes and spermatids in the central lobules, free spermatozoa in the lumen of the lobule. Ripe Primary growth, cortical alveoli, vitellogenic, and late-stage vitellogenic or hydra ted oocytes present. Postovulatory follicles (POFs) may be present. Mostly spermatozoa found in the central lobules and in the lumen of the lobule. Spent Primary growth oocytes present. Possibly some cortical alveoli oocytes present. Vitellogenic oocytes undergoing massive atresia. Few free spermatozoa in the lumen of the lobule: early stages of spermatogenesis in the peripheral lobules. served gonad and embedded in paraffin. We cut a 5.0-^m section from the sample, stained it with Harris's haematoxyUn, counterstained it with eosin (Humason, 1972), and examined it under a compound microscope to determine sex and gonad developmen- tal state. In addition, the frequency of oocyte devel- opmental stages (including atretic bodies and postovulatory follicles) was tabulated for approxi- mately 300 oocytes from each ovary by using a com- puter-driven data-acquisition software package (Optimas Corp., 1996). Gonadal development classes (Table 1) were determined by using modified classi- fication schemes developed from West (1990) and Wallace and Selman (1981) for females and from Hyder( 1969) for males. Reproductive seasonality was determined by ex- amining the monthly changes in gonad classes, the monthly distribution of oocyte stages, and the monthly changes in the gonadosomatic index (GSI). The GSI was calculated by the following equation: GSI = gonad weight/i whole weight -gonad weight). If whole weight was not available, it was estimated from TL. Batch fecundity was estimated by counting hydrated oocytes following the gravimetric method described by Hunter et al. (1985). The relation be- tween fish weight and batch fecundity was examined for a significant correlation by means of least-squares regression analysis (SAS Institute, Inc., 1985). Results Collections We sampled 858 vermilion snapper that ranged from 192 to 585 mm TL. Most fish (87% ) were between 201 and 325 mm TL. Relationships between lengths, be- tween lengths and weights, and between weights are seen in Table 2. Male and female data were pooled for the weight-length relationship because no significant difference was found between slopes and y-intercepts of sex-specific regression equations (analysis of covari- ance, P=0.47 and 0.08, respectively). Most fish came fi'om the recreational fishery (72=661) and ranged in length from 192 to 400 mm TL (Fig. 2). Fish from the commercial fishery (n= 168) ranged from 225 to 585 mm TL. The mean length of recreationally caught fish (256 mm TL, SE=29) was significantly less than the mean length of commercially caught fish (324 mm TL, SE=90; t-test, P<0.001 ). The length-frequency distributions for commercially and recreationally caught fish were sig- nificantly different (r'=216, df=5, P<0.001). Fish ob- tained fi-om the trawl survey (n=27) were small and ranged in length from 205 to 253 mm TL. The mean length of males (256 mm TL, SE=4) was not significantly different from that of females (261 mm TL, SE=3; t-test, P=0.315). Males (n =392) ranged in length from 199 to 585 mm TL and females (/2=430) from 192 to 518 mm TL (Fig. 2). The overall sex ratio of males to females was 1:1.1 and was not signifi- 832 Fishery Bulletin 97(4), 1999 Table 2 Linear relation ships (i =a-¥bX) between length and len ?th. length an d weight, and weight and weight for vermi ion snapper from the eastern Gulf of Mexico. SL is standard length (mm) FL is fork len gth(mml,TLis total 1 ength (mm); WT is w hole weight (gm); GWT is gutted weight gm). n is the number offish sampled, and standard error is in parentheses. Y X n a 6 r2 SL FL 857 7.5(0.8) 0.82 (0.003) 0.99 SL TL 869 10.5(1.0) 0.72(0.004) 0.98 FL SL 857 -5.5(1.1) 1.20 (0.005) 0.99 FL TL 854 3.4(0.6) 0.88(0.002) 0.995 TL SL 869 -8.4(1.4) 1.36(0.007) 0.98 TL FL 854 -2,6 (0.6) 1.13(0.003) 0.995 WT GWT 89 -11.8(2.3) 1.15(0.012) 0.99 log.oWT log.oTL 646 -4.60(0.084) 2.87(0.035) 0.91 logioGWT log,„TL 170 -5,01(0.05) 3.03(0.02) 0.99 0.05). Seasonal sex ratios based on three- month intervals (December-February, March-May, June-August, September-No- vember) ranged from 1:1 to 1:1.2 and were not significantly different from 1:1 (X" =0.017 to 1.045, df=l,P>0.05). Age and growth Under reflected light, alternating opaque (white) and translucent (dark) zones were evident on vermilion snapper otoliths (Fig. 1). Two readers examined a subsample of 200 otoliths, and 57% of their readings were in agreement. Most of the disagreements (82.6'^) differed by only one opaque zone. A second independent reading of the subsample by the two readers resulted in complete agree- ment (lOC^). Of 858 sectioned otoliths exam- ined, 841 OS'/f ) could be assigned ages. Of the otoliths that could be aged, measurements for marginal-increment analyses could not be made for 50 individuals (5.9'^) because of bro- ken or occluded areas along the otolith radius. Analyses of marginal-increment data for fish ages 2 to 10 suggested that opaque zones were formed once a year during the late spring to early summer (Fig. 3). During this time, the widest increments (opaque zone for- mation was imminent) and the narrowest increments (opaque zone formation was just completed ) were present. In addition, monthly median marginal increments for ages 2 to 10 had a consistent yearly pattern of high me- Hood and Johnson: Life history of Rhomboplites aurorubens 833 E E, c o E o c ra c O) CO S Ma ern 0 25-1 0,20 - 0.15 - 0 10 - 0 05 - J , Age 2 015 - 0 10 - 0 05 - 000 - 0 15 - 0 10 - 0 05 - 000 - [ 0 15 ^ 010 - 005 - 0,00 - 0 15 - 0 10 - 0 05 - 0 00 - lion s ident , » Age 7 1 ff~f^ DJFMAMJJASONDJ 0 15 ■ 0 10 - 005 - !L;_T ; Age 3 \ •» ' 3 J F M A MJ JASONDJ . Age 8 [ )J FMAMJ J ASONDJ 0 15 ■ 0 10 - 0 05 - 1 . i-r^ : J F M A MJJASONDJ Age 9 DJ FMAMJ J ASONDJ 0 15 - 0 10 - 0 05 - i+^+Hi+^ J J F M A MJJASONDJ Age 10 5— ^ ^-f 3JFMAMJJAS0NDJ 0 15 - O10 - 0 05 - T Age 6 DJFMAMJJASONDJ Monlti napper ages 2-10 years from the east- ified by the solid line. rginal Gulfc DJFMAMJJASONDJ Month Figure 3 increments from sectioned otoliths of vermi f Mexico. Median monthly increments are dian values from January to May and low values in July and August (Fig. 3). Ages ranged from 1 to 13 years. Most males (74%) and females (77% ) were found to be between ages 4 and 7 (Fig. 4). Initial growth of vermilion snapper was rapid, and fish attained a mean length of 211 mm TL during their second year (age 1, Table 3). However, subsequent increases in length were low (<33 mm). Likelihood ratio tests did not show a dif- ference between the male and female von Bertalanffy growth models ( x-=0.92, df=3, P>0.5 ). The estimated von Bertalanffy growth parameters (standard error) for all aged fish (n=841) were L„=298(5.0), K=0.25 (0.04), and ^o=-3.9 (0.85). Predicted lengths at age were similar to mean lengths at age (Table 3; Fig. 5); however, because of high variability in length-at-age data, the correlation coefficient for estimated growth was low (adjusted r^=0.26), suggesting that age is not a good predictor of length. Mortality Catch length-frequency data were transformed into age frequencies by using age-length keys constructed cfl 150 -^ Age (yr) Figure 4 Age distribution of male (open) and female (hatched! vermilion snapper from the eastern Gulf of Mexico. from our ages. Full recruitment into both the recre- ational and commercial fisheries occurred at age 6. Survivorship (standard error) estimates determined by using the Chapman-Robson method were 0.619 (6.0002) and 0.613(0.0001), respectively Instanta- neous mortality estimates (Z=-ln survivorship) were 834 Fishery Bulletin 97(4), 1999 Table 3 Mean empirical and predicted total lengths (mm TL) of female, male, and all vermi Mexico. Standard error is in parentheses, and n is number offish examined. lion snapper sampled from the eastern Gulf of Age Female Male All Fish Predicted n Mean empirical Range n Mean empirical Range n Mean empirical Range 1 6 214(13.1) 195-225 4 207(4.6) 202-212 11 211(10.1) 195-225 210 " 2 16 224(16.3) 200-255 5 221(13.4) 208-243 21 223(15.4) 200-255 229 3 27 249(24.7) 200-302 10 253(36.1) 225-347 37 250(27.8) 200-347 245 4 83 253(41.9) 202-518 66 245(23.8) 199-322 154 253(41.5) 199-518 257 5 124 267(60.51 192-518 91 262(41.51 215-505 225 270(60.8) 192-518 266 6 88 27.5(60.1) 205-515 76 262(29.7) 218-385 171 274(56.9) 205-515 273 7 29 263(23.7) 223-340 52 259(21.0) 225-315 85 267(43.1) 223-502 278 8 26 294(67.7) 235-500 32 270(14.8) 241-300 59 282(48.1) 235-500 283 9 13 277(21.8) 250-333 29 288(63.6) 325-585 45 293(64.4) 235-585 286 10 10 294(38.1) 267-397 13 295(69.6) 240-515 24 294(55.7) 240-515 289 11 2 28118.5) 275-287 2 313(40.3) 284-341 4 297(29.9) 275-341 291 12 2 271(0) 2 271(0) 292 13 2 284(32.51 261-307 3 304(42.1) 261-345 294 600 - V length (mm) 0 Q a ° V 1 3 B e '^ V o o m o •" 200 - ^p44iTTT^"TT"'==^ 100 - 0 1 2 3 4 5 6 7 8 9 10 11 12 13 Age (yr) Figure 5 Total length at age, mean length at age (solid line), and pre- dicted length at age (dashed line) for vermilion snapper from the eastern Gulf of Mexico. Circles = females, triangles = males, and squares = unknown sex. 0.480 for the recreational fishery and 0.489 for the commercial fishery. Reproduction Our sample contained no immature males and only 23 immature female vermilion snapper. Testes either contained tailed sperm, suggesting the potential for spawning, or contained brown bodies, suggest- ing previous spawning. The smallest male exam- ined was 199 mm TL; therefore males must reach maturity at a smaller size than do females. For females, 80% were sexually mature at age 1, and 100*^ were sexually mature by age 7. More than 909^ of the females were mature at 200 mm TL, and lOO'/f were mature at 325 mm TL. Vermilion snappers in the eastern GOM spawn from May to September. Ripe females were ob- served during this time period (with the excep- tion of June) (Fig. 6). These fish contained late- stage vitellogenic oocytes, hydrated oocytes, and postovulatory follicles (Fig. 7). Through the rest of the year, there was a progression from mostly resting ovaries during October to mostly early developing ovaries during December. Ripe males were observed throughout the year; they were most common (>60'^^ of testes) from March to September and in January (Fig. 6). For both sexes, median GSIs were low from October to April (<0.01 for females and <0.005 for males. Fig. 8). In May, GSIs increased dramatically (medians equaled 0.034 for females and 0.020 for males) and then gradually decreased through September. Vermilion snapper batch fecundities ranged from 553.5 to 86,811 oocytes from fish measuring 210-298 mm TL (7!=27). The relationship between batch fe- cundity and length was batch fecundity = 317 x (whole weight) - 3.1624 x 10^ r~=0.55. Hood and Johnson; Life history of Rhomboplites aurorubens 835 Discussion Collections The vermilion snapper that we sampled from the eastern GOM were smaller than those col- lected during the 1980s from the western GOM. Most hook-and-line-caught fish that we sampled (88%) were between 201 and 325 mm TL and were smaller than fish collected from the Texas Flower Gardens by Zastrow (1984) and Nelson ( 1988). The fish they sampled were mostly (757f ) between 269 and 474 mm TL and 262 and 517 mm TL, respectively. Gear bias is probably not responsible for the differences in length distributions. As in our study, both Zastrow (1984) and Nelson (1988) examined fish caught principally with hook-and-line gear. Furthermore, Nelson (1988) examined the length distributions of vermilion snapper caught with different hook sizes (Mustad no. 2-no. 4/0) and noted no differences in the size offish caught. Other factors, such as depth and movement, probably cannot explain the differences be- tween the size distribution of fish caught in our study and that offish caught in the west- ern GOM. Nelson ( 1988) caught most vermil- ion snapper from depths between 60 and 90 m. We do not know the depth at which the fish we sampled were captured; however, the mean depth at which vermilion snapper were caught by the eastern GOM commercial fishery in 1995 was 80 m ( Schirripa'' ) and is within the depth range given by Nelson (1988). The presence of larger fish in the western GOM could be the result of larger fish mov- ing from the eastern to western GOM. However, this movement seems unlikely because tagging data sug- gest that vermilion snapper are residents of reefs and do not travel long distances (Beaumariage, 1964; Fable, 1980; Grimes et al., 1982). Increases in fishing pressure may have reduced the average size of fish caught by the fishery. Schirripa^ reported that the average size of fish in the GOM commercial fishery dropped from a high of 371 mm TL in 1984 to a low of 320 mm TL in 1993. Over this same time, commercial landings increased from 1.6 million pounds in 1984 to 2.5 million pounds in 1993, and recreational landings increased from 0.2 million fish in 1984 to 1.2 million fish in 1993. The sex ratio of GOM vermilion snapper is differ- ent from the sex ratio reported for snapper from the SAB. In the GOM, female-to-male sex ratios were not significantly different from 1:1 (Zastrow, 1984; Collins'^; this study), or they significantly favored 18 25 18 39 31 14 31 82 60 42 34 21 S 20 - Female I I Immature ^^ Resting ^^ Early developing lllllll Developing I I Ripe Oct Nov Dec Jan Feb Mar Apr May Jun Jul Aug Sep IVIale NWWi Resting YX/vxA Developing 1=1 Ripe illlin Spent Oct Nov Dec Jan Feb Mar Apr May Jun Jut Aug Sep Month Figure 6 Percentage of vermilion snapper from the eastern Gulf of Mexico in gonad development classes by month and by sex. males (1:1.2; Nelson, 1988). In the SAB, females sig- nificantly outnumbered males, and sex ratios ranged from 1.6:1 to 1.7:1 (Grimes and Huntsman, 1980; Cuellaretal., 1997; Zhao and McGovern, 1997). Zhao and McGovern ( 1997 ) partitioned sex ratios from the SAB by depth, fish size, sampling year, gear type, and latitude. Of these, only gear type and latitude signifi- cantly affected the sex ratio. Females were proportion- ally more common in trap and hook-and-line collections than in trawl collections. Although Zhao and McGovern ( 1997 ) were cautious about attributing decreases in the proportion of females to changes in latitude, 1:1 sex ratios reported in the GOM (Zastrow, 1984; Collins^; our study) and from Puerto Rico ( Boardman and Weiler, 1979) are consistent with this hypothesis. We did not find any significant differences in sex ratios by season; however, Zastrow (1984) and Nelson (1988) reported an increase in the proportion of males in the summer in the western GOM. Age and growth Sectioned otoliths can be used to age eastern GOM vermilion snapper and produce higher readability 836 Fishery Bulletin 97(4), 1999 c d) 3 CT O October y '^Jm '^^:^7y '°°J^ STT^ '""J* PG CA VO LV HO ABPOF ~l T PG CA VO LV HO ABPOF PG CA VO LV HO ABPOF PG CA VO LV HO ABPOF Oocyte developmental stage Figure 7 Average frequency of occurrence of oocyte development stages by month for vermilion snapper from the eastern Gulf of Mexico. PG=primary growth oocyte. CA=cortical alveolar oocyte, VO=vitellogenic oocyte, LV=late vitellogenic oocyte, HO=hydrated oocyte, AB=atretic body, and POF=postovulatory follicle. and agreement rates than scales or whole otoliths can produce. Our readability rates were high (98%) and were similar to rates reported by Zhao et al. (1997), who were able to read 96% of sectioned otoliths from SAB vermilion snapper. The use of whole otoliths has produced mixed results. Barber ( 1989) had a high agreement rate between readings (84%) and was able to count as many as 26 zones. However, Grimes (1978) found that whole otoliths were difficult to interpret at ages greater than 7. Scales have been used with limited success by Grimes (1978), Zastrow (1984), Nelson (1988), and Collins and Pinckney ( 1988). Agreement rates between read- ings in these studies have ranged from 44% to 85% and were lower than reported in this study ( 100% ) and by Zhao et al. ( 1997 ) ( 96% ). Grimes ( 1978 ) found that counts from scales and whole otoliths from the same fish were similar and reported an agreement rate of 75%. Marginal-increment analyses suggest that vermil- ion snapper form one opaque band per year in the late spring and summer. This pattern has also been noted for vermilion snapper captured from the SAB. Zhao et al. ( 1997 ) found that opaque zones formed in June for fish age one year and in July for fish ages two to six years. Barber (1989) attempted to use marginal-increment analyses to validate annulus formation in whole otoliths from snapper from the GOM; however, he was unable to show an annual pattern. Grimes (1978) observed that the hyaline layer was formed in November in the SAB. Because hyaline and opaque bands alternate, Grimes's ( 1978) findings imply that an opaque band is formed once a year before November Vermilion snapper are considered long-lived, slow- growing fish (Manooch, 1987). The oldest individual we aged was 13 years old, similar to the SAB maxi- mum age of 12 years reported by Zhao et al. (1997), who also used sectioned otoliths. However, our maxi- mum age was older than the scale-based maximum ages reported for the western GOM of age 7 by Zastrow (1984) and age 10 by Nelson (1988). Those using scales to age fish typically underestimate the ages of older fish (Beamish and McFarlane, 1987). As fish approach their asymptotic size, little or no increase in fish size occurs and this is reflected as little or no increase in scale size. Therefore, rings at the scale edge of older (larger) fish become difficult Hood and Johnson: Life history of Rhomboplttes aurorubens 837 to interpret. Barber ( 1989) reported that otolith- based ages were on average three years greater than scale-based ages for GOM vermihon snap- per, although he did not directly compare otoliths and scales from the same fish. Our maximum age was less than the 26 years reported by Barber ( 1989). His older ages are surprising because he used whole otoliths to estimate age. Grimes (1978) noted that one disadvantage to aging ver- milion snapper by using whole otoliths is that rings are difficult to interpret in fish older than age 7. Although we did not compare whole otoliths to sectioned otoliths of vermilion snap- per, for gag (Mycteroperca microlepis), annuli at the edge of the otolith are more readily observed in sectioned otoliths than in whole otoliths (Collins et al., 1987; Hood and Schlieder, 1992). Empirical mean lengths and predicted lengths at age were less for our vermilion snapper than for snapper from the western GOM (Table 4; Fig. 9). At age 1, the mean length of fish from our study (211 mm TL) is similar to that offish from studies by Zastrow ( 1984; 207 mm TL) and Nelson (1988; 207 mm TL); however, by age 4, the mean length in our study (253 mm TL) is at least 100 mm less than mean lengths in their stud- ies (353 and 357 mm TL , respectively; Fig. 9). If scale-based ages underestimate actual age, then mean length at age and gi'owth rates will be over- estimated (Beamish and McFarlane, 1987). How- ever, because scales generally give good estimates of age for younger fish and because the mean length of our fish at age 1 is similar to lengths given by Zastrow ( 1984) and Nelson ( 1988). differences in length could reflect differences in growth as opposed to bias from aging structures. Our estimates of empirical mean lengths and pre- dicted lengths at age, which were lower than those estimates for vermilion snapper from the western GOM. could also be the product of changes in fishing pressure. Schirripa"* summarized GOM landings data and noted that annual commercial and recreational landings have increased 3- to 4-fold from the mid- 1970s to the early 1990s. If larger fish are more vul- nerable to capture, then faster-growing fish within an age class will be selectively removed from the population. The result of this type of selection will be a depression of mean size at age for older age classes and an underestimation of the biologically realistic L„ (Pitcher and Hart, 1982). This pattern was noted by Zhao et al. (1997) for vermilion snap- per in the SAB. Their estimates of growth based on ages from sectioned otoliths for 1979-81 were simi- lar to growth estimates (scale-based ages ) by Grimes (1978) from the mid-1970s, when fishing pressure 0.10 - 0.09 - Female • 0 08 - 0.07 - • • 0.06 - 1 • • 0 05 - i : 0.04 - 0.03 - / 2 0.02 - •/ • 0.01 - I^^Khh-^-^-kY ^ ' 1 c/5 0 00 " riiiiiiTiTiiifitii 1 1 1 1 1 1 '^ Oct Nov Dec Jan Feb Mar Apr May Jun Jul Aug Sep 0.10 - 0.09 - Male 0.08 - 0.07 - • • 0.06 - • 0.05 - ! • • 0 04 - • 0.03 - A • 0.02 - * / H\ : \ 1 • 001 - ! V • — 1\. 1 — I- 1 1 1-4— T •^ 0 00 T T T T T 1 T V 1 T 1 1 Oct Nov Dec Jan Feb Mar Apr May Jun Jul Aug Sep Month Figure 8 Monthly gonosomatic indices (GSI) for female and male ver- | niilion snapper from the eastern Gulf of Mexico Median val- ues are connected by the solid line. was low. Zhao et al.'s (1997) estimates of size at age and of L . then decreased over time. Estimates of L„ dropped from 629 mm TL (1979-81) to 365 (1982- 84) and 333 mm TL (1985-93) (Table 4). They could not attribute declines in size at age or L^ to gear selectivity, sampling regime, depth, or latitude and concluded their observed changes were due to selec- tive fishing for faster-growing individuals within all age classes. Similar decreases in observed lengths at age have been reported for red porgy, a species often found in the same habitats as vermilion snapper (Grimes et al., 1982; Barans and Henry, 1984; Chester et al., 1984; Sedberry and Van Dolah, 1984; Nelson, 1988) and sought by the same fisheries (Grimes et al, 1982; Nelson, 1988; Collins and Sedberry, 1991). Average total length of GOM red porgy reported by Hood and Johnson" at age 6 years was 348 mm and was over Hood, P. B.. and A, K. Johnson. 1997. Age. growth, mortal- ity, and reproduction of red porgy iPagrus pagrus) from the east- ern Gulf of Mexico. Florida Marine Research Institute. Florida Department of Environmental Protection, 100 Eighth Avenue S.E.. St. Petersburg. Florida. Manuscript in prep. 838 Fishery Bulletin 97(4), 1999 100 mm less than the average length of 464 mm re- ported by Nelson (1988). In the SAB, the average total length of 6-year-old fish had decreased from 451 mm in 1972-74 (Manooch and Huntsman, 1977) to 363 mm in 1991-94 (Harris and McGovern, 1997). Both Harris and McGovern (1997) and Hood and Johnson^ concluded that the observed decrease in average length at age was likely due to size selective fishing. Mortality The age of full recruitment to both the recreational and commercial fisheries has increased since the early 1980s. Zastrow (1984) and Nelson (1988) re- ported the age at full recruitment to be 4 years, less than our estimate of 6 years. The reported age of full recruitment to the SAB hook-and-line fishery was 4 years before 1981 (Grimes, 1978; Huntsman et al., 1983; Zhao et al., 1997) but increased to 6 years by 1993 (Cuellar et al., 1996; Zhao et al., 1997). This increase in age of recruitment has been associated with a decrease in length at age attributed to selec- tive fishing on faster-growing members of each age group (Zhao et al., 1997). Our estimates of mortality (0.469-0.489) are within the range given by Schirripa^ for the years 1986-1995 in the GOM (0.466-0.791) and less than the estimate from the SAB headboat fishery (0.67) reported by Huntsman etal. (1983). Reproduction Our observed sizes of maturation were less than those reported by Nelson ( 1988) for the western GOM. We found that most females were sexually mature at 200 mm TL (age 1), and we did not examine any imma- ture males (the smallest male we sampled was 199 mm TL). The smallest mature female and male re- ported by Nelson (1988) were 234 mm TL and 291 mm TL, respectively. In the SAB, Collins and Pinckney (1988) found that at 160 mm TL, 60% of females and 90% of males were mature. Cuellar et al. (1996) sampled females as small as 186 mm TL and males as small as 197 mm TL and did not find any immature fish in their samples. These lengths at maturity are in contrast to the results of Grimes and Huntsman (1980) who found that SAB vermil- ion snapper matured between 186 and 324 mm TL (ages 3 and 4). There is an indication that increas- ing fishing pressure may depress vermilion snapper size and age at maturity. Zhao and McGovern ( 1997) noted a decrease in the size at maturity for males and females in the SAB from 1970 to 1992. Before 1982, 31% of males and 5% of females were mature E E. c a 2 o 0) 600 - 500 - // 400 - 300 - 200 J ^ ^ 100 - ^ 0 2 4 6 8 10 12 14 Age (yr) Figure 9 Mean empirical total length at age for vermilion snapper from the eastern and western Gulf of Mexico. Scale-based ages from the western Gulf of Mexico by Zastrow ( 1984; circle, solid line) and Nelson (1988; square). Scale-based ages from Barber (1989) from the eastern (diamond) and western (hexagon) Gulf of Mexico. Whole-otolith-based ages from Barber (1989) from the eastern (triangle) and west- ern (inverted triangle) Gulf of Mexico. Sectioned-otolith- based ages from this study (circle, dashed line). at 140 mm TL. After 1982, all males and 37% of fe- males were mature at 140 mm TL. They suggested that this change may have been caused by the in- creasing and selective fishing pressure that occurred during the 1980s. On the basis of the presence of ripe females and increases in GSI, we believe that vermilion snapper in the GOM spawn from May to September. Nelson ( 1988) and Collins'' also reported summer spawning. In our samples, ripe gonads were present from late spring to early fall but were most common from May to July. In addition, GSIs were highest during the summer months. Spawning seasonality in the GOM is similar to that reported for the SAB, where fish spawn from the late spring to early fall (Grimes and Huntsman, 1980; Cuellar et al., 1996). Boardman and Weiler (1979) found that in Puerto Rican waters, spawning takes place year round. Batch fecundity was positively correlated with fish length. Nelson ( 1988) and Collins^ also found a posi- tive relationship between batch fecundity and length for GOM fish. Their estimates were larger than those obtained in our study (61,600-392,000 and 33,550- 415,161 oocytes, respectively); however, the fish sampled were also larger ( 209-510 and 248-375 mm TL, respectively). Our estimates of batch fecundity were similar to SAB estimates reported by Cuellar Hood and Johnson: Life history of Rhomboplites aurorubens 839 Table 4 Von Bertalanffy growth parameters for vermilion adapted from Goodyear and Schirripa. ' Our study snapper from the Gulf of Mexico (GOM) and South Atlantic Bight (SAB). Table and the studies of Schirripa-i and Zhao et al. (1997) were added to the table. L„ i.s the asymptotic TL, A' is the von Bertalanffy growth coefficient, Iq is the time that TL s zero, and / is the sample size Study and area Aging structure Sex n L^(mm) K to Current study Eastern GOM otoliths Combined 841 298 0.25 -3.9 Schirripa' GOM scales and otolith s Combined 586 535 .203 -0.94 Nelson (1988) Flower Gardens, TX scales Combined 906 554 0.22 -0.30 Barber (1989) St. Petersburg. FL scales Combined 34 936 0.05 -3.37 Pensacola, FL Combined 194 344 0.16 -3.356 Galveston. TX Combined 799 561 0.10 -2.81 Port Aransas. TX Combined 270 428 0.15 -2.78 Brownsville. TX Combined 779 477 0.16 -2.32 St. Petersburg. FL otoliths Combined 1113 469 0.09 -1.16 Pensacola, FL Combined 600 511 0.08 -1.44 Galveston. TX Combined 799 554 0.08 -1.46 Port Aransas. TX Combined 270 657 0.07 -1.18 Brownsville, TX Combined 779 603 0.09 -0.87 Grimes (1978) SAB scales Combined 815 627 0.20 -0.13 Zhaoet al. (1997) 1979-81 otoliths Combined 195 629 0.202 -0.117 SAB 1982-84 Combined 265 365 0.315 -0.361 1985-93 Combined 766 333 0.271 -0.899 ' Schirripa, M. J. 1992. Marine Fisheries Service Analysis of age and growth c Southeast Fisheries Center. f vermilion snapper with Miami Laboratory. Miami an assessment of the fishery in FL CRD-91/92- the Gulf of Mexico. National et al. (1996; 4000 to 90,000 oocytes from fish 186- 340 mm TL). Summary Recent stock assessments of vermilion snapper sug- gest that vermilion snapper stocks in the GOM are overfished (Schirripa''). Our study, when compared to previous studies of vermilion snapper, suggests that size-selective mortality coupled with overfish- ing may be responsible for changes in life history parameters (decreases in the mean length at age, estimated L^, and length and age of maturation). If these changes are not taken into consideration, esti- mates of the yield-per-recruit, mortality, and spawn- ing potential ratios that use the above life history parameters in their calculations could be biased. Consequently, erroneous conclusions about the health of the stock of GOM vermilion snapper could be made in future stock assessments. Acknowledgments We thank the owners, staff, clients, and especially Capt. Ed Thompson of Hubbard's Marina for their assistance in obtaining samples from recreationally caught fish. We thank Captain's Finest Seafood, Dick's Seafood, Fishin' Inc., Holiday Seafood, and Nachman's Native Seafood for their assistance in sampling the commercial fishery. We thank Lew Bullock, Eric Eaton, Dave Harshany, Dan Merryman, Christy Meyers, Heather Patterson, Fred Stengard, and Connie Stevens for their assistance in collecting and processing samples. We thank Barbara Purich and the staff of the Pathology Laboratory, College of Veterinary Medicine, University of Florida, for their assistance in the histological preparation of gonad samples. We thank Mike Murphy for his assistance in the data analyses of age, growth, and mortality. We thank Luis Barbieri, Roy Crabtree, Mike Murphy, Dana Winkelman, and two anonymous reviewers for providing valuable editorial assistance for this pa- 840 Fishery Bulletin 97(4), 1999 per. We thank Linda Torres for her assistance in the administration of the budget for this study. This study was funded by the U. S. Department of Com- merce, National Oceanic and Atmospheric Adminis- tration, National Marine Fisheries Service's MARFIN program, award NA57FF0289. Literature cited Barans, C. A., and V. J. Henry. 1984. A description of the shelf edge groundfish habitat along the southeastern United States. N.E. Gulf Sci. 7:77-96. Barber, R. C. 1989. Age and growth of vermilion snapper {Rhomboplites aurorubens) in the northern Gulf of Mexico. M.S. thesis, Te.xas A&M Univ., College Station, TX. 87 p. Beamish, R. J., and G. A. McFarlane. 1987. Current trends in age determination methodology. In R. C. Summerfelt and G. E. Hall (eds.), Age and growth offish, p. 15-42. Iowa State Univ. Press, Ames, LA Beaumariage, D. S. 1964. Returns from the 1973 Schlitz tagging program. Fla. State Board Conserv. Tech. Ser 43, 34 p. Boardman, C, and D. H. Weiler. 1979. Aspects of the life history of three deepwater snap- pers around Puerto Rico. Proc. Gulf Caribb. Fish. Inst. 32:158-182. Cerrato, R. M. 1990. Interpretable statistical tests for growth comparisons using parameters in the von Bertalanffy equation. Can. J. Fish. Aquat. Sci. 47:1416-1426. Chester, A. J., G. R. Huntsman, P. A, Tester, and C. S. Manooch. 1984. South Atlantic Bight reef fish communities as repre- sented in hook-and-line catches. Bull. Mar. Sci. 34:267-279 Collins, M. R., and J. L. Pinckney. 1988. Size and age at maturity for vermilion snapper iRhomboplites aurorubens) (Lutjandidae) in the South At- lantic Bight. N.E. Gulf Sci. 10:51-53. Collins, M. R., and G. R. Sedberry. 1991. Status of vermilion snapper and red porgy stocks off South Carolina. Trans. Am. Fish. Soc. 120:116-120. Collins, M. R, C. W. Waltz, W. A. Roumillat, and D. L. Stubbs 1987. Contribution to the life history and reproductive bi- ology of gag, Mycteroperca microlepis iSerranidae), in the south Atlantic Bight. Fish. Bull. 85:648-653. Cuellar, N., G. R. Sedberry, and D. A. Wyanski. 1996. Reproductive seasonality, maturation, fecundity, and spawning frequency of the vermilion snapper Rhomboplites aurorubens, off the southeastern United States. Fish. Bull. 94:635-653. Fable, W. A. 1980. Tagging studies of red snapper tLutjanus campe- chanus) and vermilion snapper (Rhomboplites aurorubens) off the south Texas coast. Contrib. Mar Sci. 23:115-121, Grier, H. J. 1987. Brown bodies in the gonads of the black sea bass, Centropristis striatus. In Proc. 3rd Int. Symp. Reprod. Biol. Fish, 199 p. Grimes, C. B. 1978. Age, growth, and length-weight relationship of ver- milion snapper, Rhomboplites aurorubens. from North Carolina and South Carolina waters. Trans. Am. Fish. Soc. 107:454- 456. Grimes, C B., and G. R. Huntsman. 1980. Reproductive biology of the vermilion snapper, Rhomboplites aurorubens, from North Carolina and South Carolina. Fish. Bull. 78:137-146. Grimes, C. B., C. S. Manooch, and G. R. Huntsman. 1982. Reef and rock outcropping fishes of the outer conti- nental shelf of North Carolina and South Carolina. And ecological notes on the red porgy and vermilion snapper. Bull. Mar Sci. 32:277-289. Grimes, C. B., C. S. Manooch, G. R. Huntsman, and R. L. Dixon. 1977. Red snappers of the Carolina coast. Mar Fish. Rev. 39:12-15. Harris, P. J., and J. C. McGovern. 1997. Changes in the life history of red porgy, Pagrus pagrus, from the southeastern United States, 1972-1994. Fish. Bull. 95:732-747. Helland, I. S. 1987. On the interpretation and use of r- in regression analysis. Biometrics 43:61-69. Hood, P. B., and R. A. Schlieder. 1992. Age, growth, and reproduction of gag, Mycteroperca microlepis (Pisces: Serranidae), in the eastern Gulf of Mexico. Bull. Mar Sci. 51:337-352. Humason, G. L. 1972. Animal tissue technique. W.H. Freeman and Co., San Francisco, CA, 641 p. Hunter, J. R., N. C. Low, and R. J. H. Leong. 1985. Batch fecundity in multiple spawning fishes. In U.S. Dep. Commer, NOAA Tech. Rep. NMFS. 36, p. 63-66. Huntsman, G. R., C. H. Manooch, and C. B. Grimes. 1983. Yield per recruit models of some reef fishes of the U.S. South Atlantic Bight. Fish. Bull. 81:679-695. Hyder, M. 1969. Histological studies on the testis of Tilapia leuco- stricta and other species of the genus Tilapia (Pisces:Tele- ostei). Trans. Am. Microsc. Soc. 88:211-231. Kimura, D. K. 1980. Likelihood methods for the von Bertalanffy growth curve. Fish. Bull. 77:765-776. Manooch, C. S. 1987. Age and growth of snappers and groupers. In J J. Polovina and S. Ralston (eds. ), Tropical snappers and grou- pers biology and management. Westview Press, Inc., Boulder, CO, p. 329-373. Manooch, C. S., and G. R. Huntsman. 1977. Age, growth, and mortality of the red porgy, Pagrus pagrus (Pisces:Sparidael in North Carolina. Trans. Am. Fish. Soc. 106:26-33. Nelson, R. S. 1988. A study of the life history, ecology, and population dynamics of four sympatric reef predators (Rhomboplites aurorubens, Lutjanus campechanus. Lutjanidae; Haemulon melanurum. Haemulidae; and Pagrus pagrus, Sparidae) on the east and west Flower Garden Banks, northwestern Gulf of Mexico. Ph.D. diss., North Carolina State Uni- versity Raleigh, NC, 197 p. Optimas Corp. 1996. Optimas 5.0 user's manual. Optimas Corp., Bothell, WA. Pitcher, T. J., and P. J. B. Hart. 1982. Fisheries ecology. AVI Publishing Co., Westport, CT, 414 p. SAS Institute, Inc. 1985. SAS user's guide: statistics. SAS Institute, Inc., Gary NC, 956 p. Hood and Johnson: Life history of Rhomboplttes aurorubens 841 Sedberry, G. R., and R. F. Van Dolah. 1984. Demersal fish assemblages associated with hard bot- tom habitat in the South Atlantic Bight of the U.S.A. Environ. Biol. Fish. 11:241-258. Snedecor, G. W., and W. G. Cochran. 1971. Statistical methods. Iowa State University Press, Ames, lA, 593 p. Smith, G, B. 1976. Ecology and distribution of eastern Gulf of Mexico reef fishes. Fla. Mar Res. Publ. 19, 78 p. Smith, G. B., H. M. Austin, S. A. Bortone, R. W. Hastings, and L. H. Ogren. 1975. Fishes of the Florida Middle Ground with comments on the ecology and zoogeography. Fla. Mar. Res. Publ. 9, 14 p. Vergara, R. 1978. Lutjanidae. In vol. 3: FAO species identification sheets for fishery purposes: western central Atlantic (fish- ing area 31), unpaginated. Wallace, R. A., and K. Selman. 1981. Cellular and dynamic aspects of oocyte growth in teleosts. Am. Zool. 21:325-343. West, G. 1990. Methods of assessing ovarian development in fishes; a review. Aust. J. Mar Freshwater Res. 41:199-222. Youngs, W. D., and D. S. Robson. 1978. Estimation of population number and mortality rates. In T. Bagenal (ed.), Methods for assessment offish produc- tion in fresh waters, p. 137-164. Blackwell Scientific Publications, Oxford, UK. Zastrow, C, E. 1984. Age and growth of the red snapper, Lutjanus campe- chanus. and the vermilion snapper, Rhomboplites auro- rubens, from the northwestern Gulf of Mexico. M.S. the- sis, Texas A&M University, Galveston, TX, p. 77. Zhao, B., and J. C. McGovern. 1997. Temporal variation in sexual maturity and gear-spe- cific sex ratio of vermilion snapper, Rhomboplites aurorubens, in the South Atlantic Bight. Fish. Bull. 95:837-848. Zhao, B., J. C. McGovern, and P. J. Harris. 1997. Age, growth, and temporal change in size at age of the vermilion snapper from the South Atlantic Bight. Trans. Am. Fish. Soc. 126:181-193. 842 Abstract.-Over 22,000 Atlantic cod, Gadus morhua, were tagged with T-bar tags and released in the Gulf of Maine area in 1984-97 and 2400 recovered tags were used to interpret movement of tagged fish. Most of the releases were of adult fish made during the winter cod spawning season from aggregations found on known spawning banks. Re- captures by NAFO divisions were wejghted with an annual index of fish- ing effort to account for probability of recapture. At the division level, very little exchange between the area east of 4X and the Gulf of Maine was evi- dent. However, within the Gulf of Maine an exchange of about 15% be- tween 4X and .5Z and somewhat higher between 4X and 5Y was apparent. Cod tagged on Browns Bank and Georges Bank during the spawning season showed widespread dispersal both within their respective division and to adjacent divisions. The seasonal distri- bution of recaptures in 4X indicates aggregation for spawning followed by postspawning dispersal. The seasonal pattern for Georges Bank is less clear but there are indications of net loss to the 4X area. Distribution of recaptures from Georges Bank releases in 1994 was similar to those observed for re- leases made in 1984-85. Results of the study were consistent with results from earlier tagging experiments and dem- onstrate substantial interaction of cod from different management areas. These findings may have implications for stock assessment models and man- agement objectives. Movement of Atlantic cod, Gadus morhua, tagged in the Gulf of Maine area Joseph J. Hunt Biological Station St. Andrews, New Brunswick EGG 2X0. Canada E-mail address hunt||ia'mardfo mpogcca Wayne T. Stobo Bedford Institute of Oceanography Dartmoutti, Nova Scotia B2Y 4A2. Canada Frank Almeida Northeast Fisheries Science Center National Marine Fisheries Service, NOAA Woods Hole, Massachussetts 02543-1097 Manuscript accepted 16 November 1998. Fish. Bull. 97:842-860 ( 1999), The Atlantic cod (Gadus morhua ) is a demersal species with a wide Northwest Atlantic geographic range extending from the Gulf of Maine in the south to the Davis Strait in the north ( Scott and Scott, 1988), It is one of the most important commercially exploited groundfish species in Atlantic waters off Canada and the eastern U,S, region, and cod stocks have supported fisheries in the Gulf of Maine since the 1700s (Ser- chuk and Wigley, 1992), The Gulf of Maine area, as de- fined in this study, includes waters of the Bay of Fundy, Georges Bank to 7r'41'W, and the Scotian Shelf west of 63' 20' W, an area of approxi- mately 150,000 km-. There are four fisheries management units identi- fied for Atlantic cod in this area, based on known areas of spawning aggregation and commercial fisher- ies (Halliday et al., 1986), Geo- graphic boundaries of North Atlan- tic Fisheries Organisation (NAFO) divisions and the U.S. National Marine Fisheries Service (NMFS) and Canadian Department of Fish- eries and Oceans (DFO) unit areas ( Fig, 1 ) are used to approximate the distribution of the stocks. The present management units are di- vision 4X, subdivision 5Ze in unit areas j and m (Canada), subdivision SZw+SAB (USA) and division 5Y. The area is also divided by the In- ternational Maritime Boundary (1MB) between Canada and the United States, established in 1984, Halliday and Pinhorn (1990) have provided an extensive review of the basis for existing statistical areas used to report fisheries activities and note that unit areas were es- tablished in the early 1970s, Population characteristics of the four cod stocks have been reported by Canada (Clark 1996; Hunt and Buzeta 1996) and by the United States (NEFSC, 1994). Stock abun- dance has shown substantial varia- tion in the last fifteen years and a broad range of recruitment. In gen- eral, all the stocks have declined from the long-term average abun- dance and, with the exception of the 4X stock, appear to be at low levels of spawning stock biomass. Numerous tagging studies have been undertaken to examine move- i ment of cod in the Gulf of Maine area. Hunt and Neilson (1993) pro- vided a synopsis of tag-recapture Hunt et al.: Movement of Gadus morhua in the Gulf of Maine area 843 results from experiments conducted between 1923 and 1960. An earlier summary of cod tagging stud- ies is given in Wise ( 1963 ). Templeman ( 1962 ) pro- vided a comprehensive summary of the basis for separating cod stocks in the northwest Atlantic using meristic and results of tagging experiments and concluded "there are Browns Bank and east- ern Georges Bank stocks, probably essentially separate as spawning stocks but with some inter- migration in both directions across the Fundian Channel and some intermingling of both these stocks with those of southwestern Nova Scotia." In general, results of these early experiments in- dicate substantial movement of cod within the 4X area and to a lesser extent to and from the adja- cent 5Y and 5Ze areas. The pattern of recapture locations shows a general dispersal from the area of release and some increase in the number of cod moving with elapsed time. However, it is difficult to interpret these studies as proportional movements because landings and effort are generally not avail- able for weighting probability of recapture by area. Canada has conducted a number of tagging studies on cod in the Browns Bank, Georges Bank, and Bay of Fundy areas since the 1970s. Studies conducted between 1994 and 1996 were part of a co-operative research study involving Canada and the United States. This report provides an inter- pretation of cod movements in the Gulf of Maine area based on results of recaptures of cod tagged in the 1979-97 time period. Methods Canada has conducted stratified random bottom trawl surveys in the Gulf of Maine since 1970 and on Georges Bank since 1986. The United States has also conducted similar surveys in the spring (1968) and fall (1963) of each year. Indices of abun- dance derived from these surveys are routinely used in stock assessment models to estimate popu- lation abundance (Hunt and Buzeta, 1996), and they provide a synoptic view of cod distributions in the Gulf of Maine area during both spawning and postspawning seasons. The mean catch per tow, aggregated by ten-min latitude and longitude squares, was calculated for the U.S. spring and fall research-vessel time series. These surveys cover most of the Gulf of Maine area, compared with the Cana- dian summer survey (4X only) and Canadian winter survey (5Ze only), and therefore provide a distribu- tion pattern for cod in the entire study area. Tagging locations were determined from areas of high catch rates in research vessel surveys, as well TD 68 66 64 -46 5Zo 5Zn 70 69 — 1 — 68 67 66 65 63 Figure 1 North Atlantic Fislieries Organization (NAFO) division (4X, 5Y. 5Z) and unit areas (eg. 5Zj) boundaries in the Gulf of Maine area and cod tag release locations. The dashed lines indicates the In- ternational Maritime Boundary line between Canada and the United States. as the location and timing of historical cod spawn- ing activity. Additional tagging was completed dur- ing other times in areas of fishing activity or special interest locations. Fishing trials were used to iden- tify sites with high catch rates and at least BOVe cod composition, and new sites were selected when catch rates or cod composition declined. Typically, once an aggregation of cod was located, fishing and tagging continued at that site for 1-2 days. Areas with high 844 Fishery Bulletin 97(4), 1999 catches of skates (Raja spp.) and dogfish (Squalus acanthias) were avoided because their abrasive skin resuhed in damage to cod and poor quahty specimens for tagging. Fisheries research stern trawlers, 50-70 m in length, equipped with a research Western Ila otter trawl and with wet laboratory facilities below the trawl deck, were used for large-scale tagging opera- tions. Fishing operations consisted of 5-20 min tows at depths ranging from 40 to 70 fathoms. At the end of each tow, cod were released into a sluice way lead- ing to the below-deck wet laboratory area and trans- ferred into 1500-L holding tanks with recirculating ambient sea water Viability of each cod was assessed prior to tagging and fish in poor condition or that had apparent external injuries were discarded. Con- trol experiments to assess tag or stress induced mortality were not conducted, but observation showed that even fish that were initially inverted or dormant on the bottom of the tank returned to an apparent active and normal behavior within 15-30 minutes. Tagged fish were released through a sluice way exiting about one to two meters above the water line. Approximately 100 to 250 cod could be held, tagged, and released from each tow, depending on average fish size in a tow. Fish to be tagged were removed fi-om holding tanks by hand with dip nets and placed on a measuring board to obtain length. A tag was inserted near the leading edge of the first dorsal fin and the fish was released. Total elapsed time from start to end of the tagging operation was in most cases less than 30 sec- onds for each fish. An ad hoc representative sample was examined to verify the spawning state of repro- ductive development (Hunt, 1996) but results were not recorded. Several small-scale cooperative studies were car- ried out with commercial fishing vessels. The first of these opportunistic tagging operations was conducted aboard the U.S. FV Mary V, a longline vessel fishing out of Gloucester, MA. During routine fishing opera- tions on the northern edge of Georges Bank, cod caught that were less than the U.S. minimum legal size (49 cm 19 inches) were tagged by the vessel's crew and released. Tags were also supplied to staff of the Massachusetts Division of Marine Fisheries (MA-DMF) and the Maine Department of Marine Resources (ME-DMR) to use during trips when onboard fishery observers were present on commer- cial longliners fishing in the Nantucket Shoals (MA- DMF) and on commercial charter boats fishing the inshore waters of the Gulf of Maine (ME-DMR). Dur- ing these trips, undersize cod were also tagged and released. A small number of cod, taken as bycatch in lobster traps during the winter lobster fishing sea- son (4Xq inshore), were tagged and released by Cana- dian fishermen. All tags were of the T-bar type (FD-67 and FD-94 in 1994) applied with the Mark-II tag insertion gun and were supplied by Floy Tag and Manufacturing, Inc (4616 Union Bay Place NE, Seattle, WA 98105). Overall length of the tag was about 6.5 cm and each tag had a 3.5-cm bright yellow address sleeve and a 1.0-cm attachment T-bar. Each tag was imprinted with a unique identification number and Canadian return mail address on the vinyl sleeve. Tags used in the 1994 experiment included a second mailing address for the National Marine Fisheries Service (NMFS), Northeast Fisheries Science Center, Woods Hole, MA. Tags were provided in clips of fifty with consecutive numbering to facilitate record keeping. An extensive advertising campaign, targeting Ca- nadian and U.S. fishing enterprises and communi- ties, was completed prior to and after each of the large-scale tagging experiments. This included no- tices in local newspapers and other industry publi- cations, all-weather bilingual posters located at ma- jor fish landing and processing facilities, personal contact with industry representatives, and follow-up with individuals submitting tags. A representative of the fishing sector participated in the 1994 experi- ment to obtain firsthand knowledge of operations and, as was hoped, to convey the importance of re- turning tags to colleagues. In Canada, a nominal reward of seven dollars was offered for each tag returned. In the United States, a reward of five dollars was paid for tags returned with recapture information and two dollars for tag only returns. In addition to the reward, the finder was provided with a summary of release and recap- ture information for each tag returned. All release and recapture data were loaded to a database to facilitate analysis. Tag-recapture infor- mation was edited to eliminate obvious errors such as onshore recoveries, substantial decreases in fish length, unrealistic travel and elapsed time factors, etc. Quality of recapture data was variable, ranging from no information to a detailed record of how, when, and where ( latitude and longitude ) fish were caught, fish size at recapture, age determination material, and other anecdotal information. About 80% of recap- tures included detailed information on location and about 20'7f had length-at-recapture measurements. Analysis of recapture data was completed first at the divisional level, then at the unit area level for the Gulf of Maine and finally at 10-min latitude and longitude squares. Results were compiled to show movement from area of release to adjacent areas and, conversely, movement to an area from adjacent sites. Analysis of recapture patterns at the divisional level Hunt et al : Movement of Gadus morhua in the Gulf of Maine area 845 was used to assess exchange between management areas. Observed numbers of returns were used for this purpose because exploitation rates and commer- cial fishery landings, required for weighting, were not as readily available for areas outside of the Gulf of Maine area. All releases and recaptures in divi- sions 4W, 4V, 4T, 4R, 4S, and subarea 3, which were geographically east of the division 4X-4W boundary, were grouped for this part of the analysis; we refer to this combined geographic area as "east of 4X." Because recaptures from tagging studies are usu- ally dependent on commercial fisheries, the distri- bution and abundance of recaptures is a function of the distribution of fishing effort as well as of fish movements. In an attempt to account for the effect of fishing effort, we weighted recaptures using the commercial fishery exploitation rates for cod stocks in the area. Landings, effort, and exploitation rates of cod by unit area were obtained from the published literature (NEFSC, 1994; Clark, 1996; Hunt and Buzeta, 1996). No provision was made for potential discarding, misreporting of catches, or allocation to unit area. The extent of these problems in landing statistics are thought to be substantial in some years and areas but have not been quantified (Clark, 1996; Hunt and Buzeta, 1996). A review of both the qual- ity and quantity of effort data indicated that a low proportion of landings, particularly for fixed gear components, were represented. Therefore, direct measures of effort could not be used to make adjust- ments to returns on the basis of distribution of ef- fort. However, trends in landings by unit area and fleet composition showed a relatively stable spatial and seasonal pattern. In some years a considerable proportion of total Canadian landings from division 4X was not allocated to a specific unit area. On the basis of preliminary work by Clark ( 1996 ), an algorithm was applied to allocate this unspecified proportion to unit area. The reported annual exploitation rate for each stock was then partitioned according to annual percent of unit area landings and used as a year- and area-specific index of the probability of recapture: A„,,, = ICy(C„,,x£^,), where A = adjustment factor; C = percent of annual reported landings by management area; E = reported annual exploitation rate by management area; ^ = unit area; and V = year. Within a year, tag recaptures from areas with low landings were therefore given higher relative weight- ing than those from areas of high landings. Annual differences in exploitation rate were used to weight for between-year effects. The derived adjustment factors were standardized to the unit area and year with the lowest landings and exploitation rate. Summaries of tag recaptures within an area by area of release required standardisation to reduce the impact of large numbers of releases from some areas. For example, the large number of releases from the Browns Bank area and subsequent number of recaptures would tend to overshadow recaptures from small-scale releases. Therefore all recaptures for this part of the analysis were standardized to the equivalent number of recaptures from 1000 releases. Seasonal effects of cod movements were evaluated for the two Browns Bank and Georges Bank release sites. Recaptures from these sites were aggregated by quarter with recaptures in the first month after release excluded. Results were summarized by unit area within the release division. To evaluate temporal effects, data were partitioned into groups representing recaptures from 0-12 months, >12 months after release, and total recap- tures. The influence of size at release was also in- vestigated but initial examination showed no sub- stantial difference in either direction or distance among size groups. This may be due to the fact that most fish were greater than the L^q mature reported by Hunt ( 1996) for Georges Bank cod at the time of release and would be expected to diminish the po- tential impact of size on movement. Recapture information that included latitude and longitude was used to summarize the tag recoveries by 10-min latitude and longitude squares for the Browns and Georges Bank release sites. Recaptures in each square were weighted by the adjustment fac- tor index associated with the unit area in which the square was located. The Browns Bank and Georges Bank areas support substantial commercial fisher- ies and are thought to be centers of spawning activ- ity. Tag releases were made during the spawning season and therefore recoveries should represent movement of postspawning fish. The minimum, maximum, and average time at large after tagging was calculated for releases from each unit area. For recaptures with latitude and lon- gitude location, the straight line distance between release and recapture site was calculated to deter- mine the minimum, maximum, and average distance travelled. A small proportion of recaptures included size at the time of recapture and these were used to esti- mate individual specific growth rates. The increase in length between release and recapture was adjusted to an annual value ((increment x 365)/days at large) 846 Fishery Bulletin 97(4), 1999 for each fish and averaged by 10-cm intervals of re- lease length to give an annual growth rate. Results Average cod catches per tow, derived from U.S. spring and fall research surveys for 1982-91, are shown in Figure 2, A and B, and indicate widespread distribu- tioh throughout the area with some apparent cen- ters of aggregation associated with banks. Densities and distributions show seasonal variation between the spring and fall. In the spring, distribution (with localized areas of high density) appears to be con- tiguous from Cape Cod, across Georges Bank, to Browns Bank and then farther east and west to the Bay of Fundy. In the fall, densities are lower and there is more apparent geographic isolation. Gavaris et al. (1993) found similar results when comparing seasonal distributions in relation to the 1MB line in the Georges Bank area. The central part of the Gulf of Maine area has relatively low densities for both spring and fall seasons. Over 58,000 releases were included in our study of which about 22,300 were in the Gulf of Maine pri- mary study area. The release sites in the Gulf of Maine area, aggregated by ten-min latitude and lon- gitude squares, are shown in Figure 1 and summa- rized by date and unit area in Table 1. Releases by area ranged from about 11,000 in 4Xp (Browns Bank) to only seven in 5Yd. Total recaptures were more than 6300 with over 2400 from the Gulf of Maine. Return rate by area varied between 0% and 34% with an average of 9%. Release length ranged from about 15 cm to about 140 cm. A relatively high proportion of released fish were >80 cm in the 5Zj area. The majority of released cod were larger than the L-^ ma- turity (-43 cm; see Hunt 1996), with the exception of unit areas 4Xm and 4Xo. Observed numbers of recaptures by division of re- lease and recapture for each of "east of 4X," 4X, 5Y, and 5Z areas are shown in Figure 3. Over 9T'/( of cod recaptured from releases east of 4X were also recov- ered east of 4X and, conversely, 74% of fish recap- tured in this area originated from releases east of 4X. Further analysis of information on the area east of 4X is required but, in general, movement between this area and the Gulf of Maine appears to be lim- ited. For division 4X, 80% of recaptures from releases in 4X occurred in 4X, with 8% moving east of 4X and 12% west to 5Z and 5Y. Of the total recaptures taken in 4X, 50% were from 4X releases, 42% from 5Y re- leases, and 8% from division 5Z releases. In division 5Y, 63% of recaptures from releases in 5Y occurred in 4X and of these only 21% occurred in 5Y and 2% Average number per tow ^' A spring -10 i -26 70 69 68 67 66 65 Average number per tow ■' B fall yi *-' - .,--.» Figure 2 Mean catch per tow of cod derived from 1982-91 U.S. re- search surveys in the Gulf of Maine area: (A) spring and IB) fall. from 5Z. However, of the total recaptures in 5Y, over 80% were from 5Y releases and 17% from 4X. The high proportion of recaptures in 5Y from division 4X is probably influenced by the small numbers of re- leases and recaptures as well as the distribution of commercial fishing in the vicinity of the 4X and 5Y boundary. Clark (1996) noted that Canadian land- Hunt et al ; Movement of Gadus morhua in the Gulf of Maine area 847 100 75 c B 50 Q. 25 East of 4X By Division ^/ of recapture 4X 5Y 5Z By Division y"^ of release n L X > K s I U U J JiJL . Division Figure 3 Percent of cod tag recoveries by NAFO Division of release (solid bar) and Division of recapture (pattern bar, based on 1000 releases I for divisions 4X, 5Y, 5Z, and the area east of 4X. ings from the 5Y area near the 4X-5Y boundary are considered part of the 4X fishery and therefore were included in our 4X stock assessment. In division 5Z, 77% of cod released in 5Z were recaptured in 5Z and 21% were recaptured in 4X. Of the total recaptures in division 5Z, 81% were from 5Z releases and most of the movement out of the area occurred to the north- east (15%), towards division 4X. The annual percent exploitation rate of cod by management area (4X, 5Y, 5Z) is summarized in Fig- ure 4 for the 1980-97 time period. Reported land- ings by unit area showed considerable variation but proportions were relatively stable over time. In divi- sion 4X, unit area 4Xo accounted for about 38% of the 1980-97 total landings and unit area 5Yd for about 40% of the 5Y total. In subdivision 5Ze, unit areas 5Zg(30%) and 5Zj (40%) dominated, represent- ing the U.S. and Canadian fisheries, respectively. The derived recapture adjustment factors, standardized to a maximum of 1.00, are given in Table 2 for each year and unit area. Annual factors ranged from 0.02 to 1.00, meaning that about 100 recaptures from a unit area with high landings and exploitation were proportionally equivalent to about two recaptures from a unit area with low landings and exploitation rate. The effect of weighting tag recoveries to account for fishing effort is shown in Figure 5 for releases in the Browns Bank area. Weighted recoveries decrease ^the apparent extent of movement to 4Xo and 5Zj, because of their relatively high landings and exploita- 0 808284868890929496 Year Figure 4 Reported percent exploitation rate of cod by the commercial fishery in management areas 4X, 5Y, and .5Z for 1980-97. tion rates, and increases the contribution of 4Xp recov- eries. Similar effects are seen for other unit areas. Table 3 provides a detailed summary of recaptures by unit area of release and recapture partitioned into <12 months elapsed time from release and the total number of recaptures. The observed number of tag returns, the number adjusted by unit area landings, and the percent of total adjusted number are shown. The percent of recaptures reported is based on only those released and recaptured in the Gulf of Maine 848 Fishery Bulletin 97(4), 1999 area. Table 3 summarizes results by area of release and shows the pattern of movement away from the release site (or no movement when the area of release=area recapture). Table 3 shows the pattern of movement into an area from standardized releases in adjacent areas and is organized so that the two aspects of tag recoveries appear on the same hori- zontal line. For example, in the second row of data, results show 31.4% of cod released in 4Xm were re- caf)tured in 4Xo, whereas only 15.5% of cod recap- tured in 4Xm originated from 4Xo releases. About 63% of the cod tagged in 4Xm appeared to remain in the area although there was some move- ment to the west into adjacent areas. Over 76% of cod recaptured in 4Xm originated from 4Xm releases. At the divisional level, movement in relation to the division 4X/4W boundary is small (Fig. 3); however, exchange between divisions does occur and is con- centrated in the 4Xm, 4Xn, and 4Xo areas. No releases were made in 4Xn but about 80 recap- tures were taken in the area. Most of these recap- tures were from the Browns Bank (46% ) and Georges Bank (27%) releases. Additional recaptures origi- nated in the Bay of Fundy area and about 4% from the area east of 4X. Results for the 4Xo area indicate a contrast be- tween the two analyses of recaptures. Close to 100% of recaptures of cod released in the 4Xo area were Table 1 Summary of cod tag releases and recap ures in NAFO divisions by release date. * indicates release from commercial fishing | vessels. East of Area released 4X 4Xin 4Xo 4Xp 4Xq 4Xr 4Xs 5Yb 5Yc 5Yd 5Zj 5Zm 5Zo Date 1978-81 Feb 80 May 79 Feb 84 Dec 94* Jun79 Aug 84 Jul 85 Jun 94* Jul 94* Feb 84 Mar 94 Jul 94* N 36,338 808 694 3,980 50 141 67 350 27 7 1988 843 100 Length range (cm) 15-135 13-92 21-87 3.3-140 36-64 22-75 38-100 ,33-112 34-47 36-43 16-136 31-130 3,5-48 Mean length (cml 48.8 47.9 41.9 71.9 47.4 36.7 57.1 59.9 41.1 40.3 61.2 69.9 42.4 Date Mar 84 May 94* Jul 85 Dec 85 Mar 85 Jun 94* N 786 4 216 73 500 266 Length range 40-126 38-57 28-103 26-64 33-139 30-50 Mean length 66.3 43.5 52.9 42 72.1 44.4 Date Feb 85 Dec 95* Jul 85 Mar 94 N 1.621 9 1399 3608 Length range 38-142 38-72 26-94 30-144 Mean length 79.8 48.1 52.6 73.8 Date Mar 85 May 96* Nov 85 N 4363 58 281 Length range 32-140 39-85 28-87 Mean length 72.9 58.8 48.9 Date Nov 96* Jan 85 N 17 4 Length range 39-72 36-56 Mean length 53.8 42.3 Date Jul 94* N 21 Length range 36-47 Mean length 40.3 Total released 36,338 808 694 10,7.50 138 357 1845 350 27 7 6096 1109 100 Total recaptures 3883 100 237 1346 1 57 139 49 0 0 476 29 0 Percent recaptures 10.7 12.4 ,34.1 12.5 0.7 16.0 7.5 14.0 0.0 0.0 7.8 2.6 0.0 Hunt et al : Movement of Gadus morhiia in the Gulf of Maine area 849 recaptured in 4Xo. Cod tagged in 4Xo were from an inshore area and may have been part of a local resi- dent population. About 907^ of tags recovered in 4Xo were from fish released in 4Xo and additional small contributions were fi-om other Gulf of Maine locations. The largest number of releases was ft'om the Browns Bank area and results indicate a wddespread dispersal within the Gulf of Maine. The majority of recaptures were from division 4X; substantial numbers of cod moved to the inner Bay of Fundy and smaller num- bers moved east to 4Xn and 4Xo. About 'iy/c of cod released in 4Xp were recaptured in the same area. Movement into Divisions 5Y and 5Z accounted for about 157f of the total releases. Of cod recaptured in 4Xp, about SO'/f originated from the same area, 21^/c from 5Yb, and about Vi% originated from the Georges Bank area. Differences between immigi'ation and emi- gration distributions may be due to seasonal patterns and commercial fishing operations. The Browns Bank area has been closed to commercial fishing during the winter spawning season and there is a low probabil- ity of capture during this time. Cod recaptured dur- ing the summer-fall fishing season may represent postspawning dispersal patterns. In 4Xq, all three of the recaptures were taken in 4Xo. However, cod caught in 4Xq appeared to origi- nate from diverse locations including 39''r from 5Yb, 28% from Browns Bank, 20% from the inner Bay of Fundy, and 10% from Georges Bank. 50 40 - _ 30 20 10 a I Adjusted for fistiing effort / Unadjusted II a^ ^^j^ ij- in u^ in Unit area of recapture Figure 5 Comparison of percent recaptures by unit areas for Browns Bank releases using observed and effort-adjusted tag recoveries. Unit area 4Xs and 4Xr represent the inner Bay of Fundy region. Movement of cod appears to occur into and out of the region as well as across the Bay from the Nova Scotia to New Brunswick sides. Over 75% of recaptured cod released in these areas were re- Table 2 Annual adjust ment factors by year and area used to prorate recoveries for probability of recapture, standardized to 1.00 for the area an d year (5Zn, 19871 with the lowest Ian dings and ex jloitation rate. Area 1980 1981 1982 1983 1984 1985 1986 1987 1988 1989 1990 1991 1992 1993 1994 1995 1996 1997 Mean 4Xm 0.13 0.13 0.10 0.16 0.20 0.10 0.10 0.10 0.13 0.24 0.17 0.14 0.14 0.14 0.19 0.71 0.84 0.67 0,24 4Xn 0.21 0.28 0.18 0.32 0.20 0.25 0.17 0.17 0.18 0.32 0.41 0.22 0.12 0.12 0,28 0.30 0,47 0.36 0,25 4Xo 0.07 0.05 0.05 0.06 0.07 0.06 0.06 0.06 0.04 0.06 0.04 0.04 0.02 0.04 0,04 0.12 0.14 0.14 0,06 4Xp 0.44 0.33 0.25 0.22 0.27 0.29 0.54 0.24 0.15 0.12 0.18 0.14 0.15 0.14 0,14 0.13 0.25 0.23 0,23 4Xq 0.19 0.24 0.14 0.16 0.15 0.12 0.11 0.12 0.12 0.17 0.12 0,13 0.08 0.10 0,11 0.16 0,24 0.17 0.15 4Xr 0.15 0.12 0.11 0.13 0.12 0.11 0.13 0.23 0.21 0.24 0.16 0,09 0.07 0.08 0,10 0.21 0,24 0.14 0.15 4Xs 0.30 0.30 0.27 0.34 0.36 0.23 0.23 0.28 0.28 0.46 0.23 0,15 0.16 0.15 0,21 0.50 0.43 0.28 0.29 5Zg 0.04 0.05 0.04 0.05 0.05 0.04 0.07 0.08 0.05 0.05 0.05 0.04 0.06 0.04 0.03 0.06 0,11 0.08 0.06 .5Zh 0.14 0.16 0.16 0.12 0.13 0.12 0.25 0.16 0.10 0.15 0.20 0.13 0.14 0.08 0,08 0.17 0.33 0.24 0.16 5Zj 0.08 0.06 0.04 0.04 0.05 0.04 0.05 0.04 0.03 0.05 0.04 0.03 0.02 0.02 0,03 0.20 0.23 0.12 0.06 5Zm 0.33 0.26 0.23 0.23 0.10 0.10 0.13 0.22 0.10 0.13 0.15 0.11 0.12 0,08 0,11 0.32 0.55 0.62 0.22 5Zn 0.39 0.40 0.71 0.97 0.28 0.38 0.79 1.00 0.57 0.53 0.32 0.21 0.34 0.19 0,15 0.33 0,64 0.46 0.48 5Zo 0.49 0.76 0.44 0.38 0.29 0.22 0.32 0.78 0.34 0.40 0,32 0.27 0,46 0.33 0.18 0.41 0,79 0.57 0.43 5Yl3 0,61 0.76 0.41 0.31 0.30 0.15 0.19 0.13 0.33 0.44 0.28 0.06 0,06 0.07 0.07 0,07 0.07 0,07 0.24 5Yc 0.15 0.15 0.10 0.11 0.10 0.08 0.06 0.06 0.13 0.07 0.07 0.08 0.08 0.09 0.09 0.09 0.09 0,09 0,09 5Yd 0.02 0.02 0.03 0.03 0.02 0.02 0.02 0.02 0.02 0.02 0.02 0.02 0.02 0.02 0,02 0.02 0.02 0.02 0.02 5Ye 0.02 0.02 0.02 0.02 0.03 0.03 0.04 0.05 0.04 0.05 0.05 0.04 0.04 0,04 0,04 0.04 0.04 0.04 0.04 .5Yf 0.12 0.09 0.07 0.09 0.08 0.05 0.06 0.05 0.12 0.09 0.07 0.07 0.07 0.08 0.08 0,08 0.08 0,08 0.08 850 Fishery Bulletin 97(4), 1999 T1 C ^^ CO a m 1 o o c. X c n. o H OJ \~ 1 \/ a. ,2 a: 3 cu X) OJ e OJ 1 en c T3 'n D. « 3 E ^ 3 ■.V OJ m -a P:: rt OJ s; 3 e 3 CO DS M CCC^'XJlOOO^COt^C^QO^OC^CC^ c^cO'Hr-icooocccD CO CTJ O l> ^ C^ o 00 IC 02 o a> ID CD ^ lO ■-I •-' C<1 i-H 00 o CO 00 iC Tj< C£> 00 lO O) 03 d d LC lO t^OCOt^'^OltDOCDOOCO 00 c- W 0^ O] COin'*Tf^lM— ICNCMCNO CO C<1 t-^ (M Oi O 05 o lO ^ in •* CD Tf CO CO CO 00 O c~ .-H Tf 00 C5 to 00 oa CO in T-H CO CO CD CD ^ CD o c~ ■* o 00 rt CD I— t .—I C^ 00 CO C<1 ■rrLoaoica^cMO^'— ii— iiolt: oi m OCO'MOO^^C£!'-'000 CC CM C^ CO CD 1-: Tt d O CD CO oi 1 CO ■-< 00 UO ?XXXXXXX »->->->->->-s:NtS!SitsisiX X o X 9- C -* 'T CM 05 u CO f-» CO I— 1 cu CD CO cu T3 X X3 -Q 3 LO c^ •-I lO 00 IC IC CM CD ■ ■ ■ ' CM O^ ^ ^ O — d d d d X C^ LC in in CD CM o d d -^ ^ c^i d d o lO CM 1-1 00 O CMOiCM^CMCMCMt^ ^C-^t^OlOOOiCD IC CO lO CO — ^ O -HCDOOOJCDCMCMrt r1 CO --C --H rt CD -^ CD CM CD CO CO --H ^ O •-» •-I ^ CM O ^ .-H t- CD ■ '^ d '^ c: o Ci. o* X X X X Tt ■* tT Tt e X IN X X Tf m LO 00 CO ^-* ._C C O Q-CTt- v;-D OT3(— bC3 bJ}X--^£ C O O [?X XXXXXX><>->^>->->-tSItSl N^N S) SI X uOTtTf-3'TrTT^'^ir^iommmicuomuoioioiC'^ X a X Hunt et al Movement of Gadus morhua in the Gulf of Maine area 851 ^ V 3 C 'S c o u ^ ." ^ O 05 C~ (C O ,-H 00 OJ t> X (N IT- 00 O O) 01 Ol O Ji E OS c PS ?; t-- m CO o 05 « CO CD -^ CO o CO o CM o o to O CO 00 O) < = ^ a ^ t. 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N N Tf lO UO lO s IS] U3 Hunt et al.: Movement of Gadus morhua in the Gulf of Maine area 853 40 20 0 20 40 60 covered there and between 15% and 20% of recaptures were from releases in the 5Yb area. Cod tagged in unit area 4Xs show more widespread dis- persal than those from 4Xr. The small number of releases from division 5Y precluded detailed inter- pretation of movement. However, there is an indication of association with the 4X area, as well as movement from the Browns Bank area into 5Y. both of which were not unexpected because the majority of releases were near the northeastern border of 5Y. Results, therefore, probably do not reflect those for the remainder of the 5Y area. The pattern observed for Georges Bank is similar to that seen on Browns Bank. Dispersal of postspawning cod appears to be substantial in both a north and south direction. Interaction between the Browns Bank and inshore area and Georges Bank accounted for most of the movement of tagged cod. The interchange was bidirectional, but more cod appeared to move from Georges Bank to Browns Bank (about 30%) compared with about 3% of Browns Bank cod moving to Georges Bank. Relatively small numbers of cod released in the 5Zj and 5Zm area were recaptured in areas further to the south or west. However, Hunt and Buzeta ( 1996) noted that the U.S. com- mercial fishery, and therefore the prob- ability of recovering tags from the 1994 and later releases in the area west of the 1MB, has been substantially reduced with the introduction of closed areas in 1995. Information for the Browns Bank and Georges Bank areas is shown in Figure 6 for adjusted recap- tures in the area and from the area. Preliminary com- parison of recoveries in the first twelve months and total recoveries showed little variation in the distri- bution or proportions ( Table 3 ) and therefore total re- coveries were used. Dispersal from the area of release appears to be substantial for both the 4Xp and 5Zj ar- eas. However, a high proportion of fish recaptured in the two areas was released fi-om the same area. The two patterns may represent postspawning dispersal (top panel for each area) and summer distribution or year-round residents (bottom panel for each area). The seasonal distribution of recaptures from the Browns Bank and Georges Bank areas is shown in Figure 7. Recaptures were combined by quarter and I III Unit area of recapture for cod released In 4Xp n ° Unit area of release for cod recaptured In 4Xp in Lo un 60 40 20 0 -20 -40 -60 -80 Unit area of recapture for cod released in 5Zj -.|l-- I Unit area of release for cod recaptured in 5Zj E X Unit area Summary Georges B Figure 6 of adjusted cod tag recoveries for the Browns Bank l4Xpl and ank (5Zj) areas, aggregated by unit area. unit area (by division for recaptures outside the di- vision of release). In the Browns Bank area, seasonal distribution appears to be consistent with aggrega- tion in the spawning season (first quarter) and with progressive dispersal to adjacent areas in the remain- der of the year. This pattern may be repeated on an annual basis with return of migrants during the spawning season. Seasonal trends in the Georges Bank area are more difficult to interpret and the in- fluence of fishing season may be a factor, compared with trends for Browns Bank, because closure to com- mercial groundfishing in the first half of the year limits potential recoveries during the spawning sea- son. However, the recapture pattern for cod recov- ered in division 4X suggests that the migrants do not return to Georges Bank because the proportion remains relatively constant over the first to third quarters and increases in the fourth quarter. 854 Fishery Bulletin 97(4), 1999 Figures 8 and 9 provide information on the spatial dispersal of cod tagged in the Browns Bank and Georges Bank areas with adjusted recoveries aggre- gated by 10-min squares and expressed on a per thou- sand basis. Most recoveries were made near the re- spective release site. Wlien movement occurred, a general pattern of dispersal from the release site with a gradient from the tagging site to the edges of the distribution was evident. Intermediate areas of ag- gregation were consistent with the geographic dis- tribution of fisheries effort and associated probabil- ity of recapture. Interchange occurred between re- lease site areas, but the main vector of movement from both sites was towards the northeast. As a re- sult, most long distance recoveries were made in the 1000 800 600 400 200 Released in 4Xp / la jj] iJll J^ JkikD 4Xm 4Xn 4Xo 4Xp 4Xq 4Xr 4Xs 5Y 5Z 800 600 400 200 Released in 5Zj,m 4th quarter 0 - I 1st quarter m d 5Zg 5Zh 5Zj 5Zm 5Zn 5Zo 4X 5Y Recapture area by quarter Figure 7 Quarterly distribution of cod tag recoveries from the Browns Bank (4Xpl and Georges Bank (5Zji release areas, aggregated by unit area. southwestern Nova Scotia and Bay of Fundy inshore areas, although there was limited movement into the northwestern Gulf of Maine or western Georges Bank areas. The two time periods associated with Georges Bank tagging are compared in Figure 10 for the 1994 and 1984—85 experiments. The two distributions are simi- lar and both show the majority of cod remaining near the release area; more of the long distance disper- sion occurred into the 4X area than to the west into the inner Gulf of Maine. Similarity in the results from the two time periods indicates some consistency in the northeast vector of movement. However, a higher proportion of cod were recaptured in 4X from the 1994 experiment (559f ) compared with the earlier releases (42%). As noted above, the reduced U.S. commercial fishery in the area west of the 1MB since 1995 also reduced the probability of tag recaptures from the 1994 experiment in this area. The distribution of recaptures and elapsed time for the February and March releases for unit areas 4Xp (Browns Bank) and 5Zj (Georges Bank) are shown in Figure 1 1 . The majority of returns in both areas occurred in the first 24 months after release, and the maximum time at large was 87 months (Table 4), but the overall average of time at large was less than 8 months. Many fish were recaptured in the first month after release, probably before postspawning dispersion had occurred. Maximum distance trav- elled was over 620 km and the mean distanced trav- elled was 60 km. Substantial numbers of cod moved hundreds of kilometres, indicating that cod can sur- vive the stress of tagging and are capable of long dis- tance movements. A strong seasonal pattern, consis- tent with the seasonal nature of the commercial fish- ery, is evident. The fishery on Browns Bank is typi- cally year round, with peaks late spring and sum- mer (Clark, 1996), and on Georges Bank, the fishery typically opens in June (Hunt and Buzeta 1996). The seasonal peaks in recaptures (4—6, 16-18, etc. months after February releases! coincides with the commer- cial fishing activity. Length at the time of recapture was available for a relatively small proportion of recaptures. For these samples, the net increase in length and elapsed time was used to calculate a specific growth rate, ex- pressed as annual rate, for each unit area. Results are given in Table 5 for 10-cm length-at-release in- tervals but sample size was small for all but the 4Xp, 4Xs, and 5Zj areas. The calculated annual growth rates indicate differences between areas and follow an expected pattern of decrease as the length at re- lease increases. Of these three areas, the Georges Bank area had the fastest growth with an annual increase of about 19 cm for cod released as 40-cm Hunt et al.: Movement of Gadus morhua in the Gulf of Maine area 855 individuals. Estimates of growth rates from this study are consistent with the more ex- tensive study on cod growth completed by Shackelletal. (1997). Discussion Fisheries management objectives for either conservation or yield from a defined manage- ment unit are usually based on a closed sys- tem approach that requires that emigration and immigration of fish from adjacent stocks be sufficiently low so as to avoid confounding stock-status evaluations. Gulland's (1983) operational definition of a stock also consid- ered interchange between stocks and he sug- gested that it can be ignored if conclusions on the population dynamics of the stock remain valid. Therefore, two problems require evalu- ation in assessing stock definitions: 1) Is there evidence of interchange offish at a geographic scale larger than the one used to define a stock boundary? and 2) If interchange does occur, what are the impacts on population dynamics and on the assumptions used in models to es- timate population parameters? Tagging experiments can address the first of the two questions provided that the study logistics have not compromised the results and that both ongoing fisheries removals and manage- ment activities during the recapture period do not invalidate interpretation of the results. Capture, tag- ging, and release of cod with bottom trawl methods and T-bar tags can generate a large number of re- leases and, according to a relatively high recapture rate and extended time at large, appear to result in nominal tag- or stress-induced mortality. During the 1994 experiment, cod tagged from one tow were re- captured in a subsequent tow, either the same day or several days later and thus may indicate that tagged cod resume a normal schooling pattern shortly after release. Efficiency of the tagging operation is also a contributing factor as is the time of year, depth, and condition of tagged fish. Clay et al. (1989) re- ported on tagging experiments under winter condi- tions and concluded that cod at this time of year have a better chance of surviving the tagging process. Tag loss was not examined in our study but Saunders et al. (1990), in a study of sablefish tagged with tags similar to those used in our study, concluded that tag loss could be as high as 10% in the first year and 2% per year thereafter. Less than 10% of cod tagged with T-bar tags for identification in a live-fish hold- ing facility experienced tag loss in the first week, al- Figure 8 Distribution of adjusted cod tag recoveries from releases in the Browns Bank (4Xp) area, aggregated by 10-min latitude and longitude squares. though losses were higher after biweekly netting and sampling for length and weight observation (Nelson' ). In the present study, the influence of high annual mortality rates (Z), failure to report recaptures, and the effectiveness of advertisements and rewards, the impact of fishery controls, and the potential bias be- tween Canada and U.S. reporting rates may be more substantial than factors such as tag loss. However, these variables are difficult to quantify. All cod stocks in the study area have been sub- jected to very high exploitation rates in the last de- cade. For example. Hunt and Buzeta ( 1996 ) reported rates of over 40% in the 1990s for the Georges Bank cod stock. A similar capture rate applied to the tagged population [N] could result in an apparent rate [R] of over 20% returns after accounting for tagging mortality (-20% ) and tag loss (-10%) and assuming full reporting \R = 0.4(7V - A^(0.2 + 0.1) = 0.287V]. Observed return rates were less than half those for the Georges Bank and Browns Bank areas and a lack of reporting could be a contributing factor. Anecdotal information from field staff working in fishing ports ' Nelson, C. 1996. Department of Fisheries and Oceans, St. Andrews, New Brunswick, Canada. Personal, commun. 856 Fishery Bulletin 97(4), 1999 indicates that some fishermen with tags did not re- turn them because of resistance to identifying fish- ing locations and because of preconceived ideas and concerns about the significance of the release and recapture site. The extent of a lack of reporting is difficult to quantify and its effect may not introduce bias if nonreported tags have the same geographic distribution as reported tags. Advertizing campaigns for the 1994 experiment were much more extensive than those used in the early 1980s. Printed notices covered a larger audi- ence, the experiment was discussed with industry representatives and conducted with their support, interim results were presented, and a more active role was taken by field staff to collect recovered tags. However, the return rate was higher in the 1980s. For example, the return rate in the first 24 months for the 1984—85 Georges Bank experiments was about 12.2% compared with 5.5% for the 1994 experiment. Fisheries management controls may play a more substantial role in determining the number of recap- tures. Controls may include area closures, restricted seasons, and catch limitations. The Browns Bank area has been closed to fishing during the winter Ftecaptures (per rriile) Figure 9 Distribution of adju.sted cod tag recoveries from releases in the Georges Bank (5Zj) area, aggregated by lO-min latitude and longitude squares. spawning season for many years. Canadian commer- cial fishing on Georges Bank has been closed between January and May since 1994, and the United States has an extensive closed area covering most of the northeast part of the Bank. These restrictions sub- stantially reduce the possibility of recaptures dur- ing the spawning season. The establishment of the International Maritime Boundary in 1984 between the United States and Canada and the associated opinions and concerns of the fishing industry have undoubtedly introduced bias in the tag return rate and perhaps in the re- ported recovery location. It might be assumed that the direction of the bias would tend to support the concept of national ownership of the resource by the fishing industry and that tags recovered at times or locations that appeared to contradict this opinion would be withheld or discarded. Comparison of tag recoveries from the 1984—85 releases with those from the 1994 releases does not show substantial differ- ences in distribution offish. The first period includes the time when response to international boundary change would have been most sensitive and pro- nounced. The second period, more than ten years after the change, might be expected to have reduced concerns by fishermen on tag location. However, the impact of nonreporting bias would be of most concern in evaluating the pro- 46 portional spatial distribution of tag recover- ies. Bias in reported recapture location is of less concern because fishing activities are monitored and there is limited opportunity to misreport the area of operations. Total allowable catches and landings of cod from the Georges Bank area in 1994 declined to less than 50% of the recent ten-yr average and are expected to remain at low levels for a number of years (Hunt and Buzeta, 1996). The impact of reduced catches is accounted for, to some extent, by reduced annual exploitation and its impact on tag-recapture adjustment. However, other changes in the fishery, such as shifts in gear sectors and spatial redirec- tion of effort, may offset this adjustment. Even with the need, discussed above, to qualify the interpretation of tagging data, we believe the results of the present study clearly indicate that interchange between the 5Zj,m and 4X cod management units occurs with a net movement from 5Zj,m to 4X. Cod released on Georges Bank exhibited movement outside the 5Z management area onto Browns Bank and the inshore areas of Nova Scotia and the Bay of Fundy. Cod released on Browns Bank also moved to the inshore area of Nova Scotia Hunt et al,: Movement of Gadus morhua in the Gulf of Maine area 857 as well as onto Georges Bank. These results correspond to those of historical studies, al- though conclusions from the earlier studies may be more limited in scope owing to the smaller number of released fish and the lower intensity of commercial fisheries at the time. In his review of tagging results from the late 1890s to the 1960s in the Gulf of Maine, Wise (1963) concluded that cod of the offshore ar- eas of Browns and Georges Banks were closely related to fish of the southwestern Nova Scotia area. Wise and Jensen ( 1960 ) had earlier con- cluded that the eastern Georges Bank cod population mixed little with the more west- ern or southern components and interacted primarily with the southwestern Nova Scotia area. Templeman (1962) in summarizing cod tagging information for the northwest Atlan- tic concluded that there were discrete spawn- ing stocks on the eastern part of Georges Bank and Browns Bank and mixing in both direc- tions across the Fundian Channel as well as to inshore areas. Results of tagging studies completed in 1969 on Browns Bank and in 1972 from inshore areas of Nova Scotia were reported by Halliday ( 1973) and he concluded that there was a separation of inshore and off- shore stocks. McKenzie (1956) and McCracken (1956) tagged cod in the inshore area of south- west Nova Scotia and observed limited inter- change with the offshore banks and the Bay of Fundy. They concluded that the resident inshore cod stocks in the Bay of Fundy and southwestern Nova Scotia were relatively sta- tionary. They also explained the offshore re- captures of cod tagged in the inshore area as evidence that cod from more migratory stocks could be present in inshore areas at certain times of the year. Hunt and Buzeta (1989) provide details of the basis for partitioning the northeast part of Georges Bank (5Zj,m) as a separate Cana- dian management unit distinct from the re- mainder of the 5Zw-i-SA6 area. They concluded that spatial distribution from tagging studies and other biological characteristics were suf- ficiently distinct to define the 5Zj.m area as a management unit that could be expected to benefit from management controls. However, the 5Zj,m area is partitioned by the 1MB so that commercial fisheries by Canada and the United States are limited to their respective sides. Gavaris et al. (1993) conducted an an- alysis of cod movement in relation to the 1MB using commercial catches and research sur- Recaptures (per mille) from 1 984-85 releases -'^Kj^-''^ 70 -T^ — ' r 69 68 67 — r 66 — r 65 64 63 Figure 10 Di.stribution of adjusted cod tag recoveries from releases in the Georges Bank (5Zj) area, aggregated by 10-min latitude and longitude squares: (Al 1994 releases; (Bi 1984-85 releases 858 Fishery Bulletin 97(4), 1999 Table 4 Summary by unit area of release of months after release to recapture and distance travelled I km I. Months after release Distance travelled Area of release Minimum Maximum Average Minimum Maximum Average 4Xm 0 87 12.0 1.12 385 64.5 4Xo 0 25 2.4 0 448 7.2 4Xp 0 62 12.2 2 626 128.6 4Xq 7 7 7 24 24 24.0 4Xr 0 28 4.4 0 235 25.4 4Xs 0 38 8.5 0 385 75.8 5Yb 0 31 7.0 2 221 66.5 5Zj 0 65 10.1 2 367 62.3 5Zm 3 16 5.3 22 265 93.4 Average 0 39.9 7.6 5.9 328.6 61.0 4Xp Releases 30 >. 0 lliillll-Jli-ii.j-.IL-«i-.ill-.— ■!■■-■ C 0 5 II 16 21 27 34 41 58 5Zj Releases Jll I iL.lliI I......J....JJ. !.... 0 5 11 16 21 27 36 42 50 59 Elapsed months since release Figure 11 Summary of adjusted cod tag recoveries for releases in the Browns Bank (4Xp) and Georges Bank (5Zj ) areas by elapsed time from release to recovery, in months. vey indices. They concluded that extensive move- ment does occur with a strong seasonal pattern in which almost 100% of the biomass is on the Cana- dian side in the fall-winter period and about 65% in the spring-summer period. Results of the present study support the occurrence of seasonal movement but also indicate that it is not a closed system. There is evidence of immigration and emigration and an apparent net loss from the Georges Bank area to the Browns Bank and division 4X area. In division 4X, two substocks (Bay of Fundy and Scotian Shelf) are already assumed (Clark, 1996) with distinct growth characteristics. However, tag- ging results show interchange between the Bay of Fundy, the inshore areas of southern Nova Scotia, and Browns and Georges Banks. The greatest ex- change occurs between the offshore Browns Bank area and the inshore areas and the Bay of Fundy. Our results indicate that the extent of movement by cod in the Gulf of Maine area is substantial and that it crosses the present stock boundaries between 4X, 5Y, and 5Z. The interchange appears to be pri- marily between the eastern part of Georges Bank (the Canadian management unit ) and 4X. Although movement between 4X and 5Y was observed, the number of releases and recaptures was small and the tagging location was close to the boundary be- tween these areas. Exchange between 5Z and 5Y also occurs but at a relatively low rate. Investigation of the potential impact of cod move- ment between stocks on models used for population estimates will require further work. Interpretation of diagnostics for population models such as ADAPT (Gavaris, 1988) from a stock concept might give Hunt et al.: Movement of Gadus morhua in the Gulf of Maine area 859 Table 5 Growth rate of cod estimated from net increase in length and elapsed t me, expressed as an annual rate . Number of observations | are shown in parentheses. Release length (cmi 4Xm 4Xo 4Xp 4Xr 4Xs 5Yb 5Zj 20- -29 12.5(2) 12,0(4) 30- -39 13.4(7) 14.9(6) 17.7(8) 26.6(4) 40- -49 7.9(7) 8.9(8) 12.4(7) 15.8(15) 33.8(5) 18.9(16) 50- -59 10.7(22) 8.0(4) 9.2 (28) 2.6(3) 15.8(16) 30.7(3) 16.9(12) 60- -69 5.1(4) 7.5(66) 18.0(2) 10.8(18) 18.9(9) 12.2(45) 70- -79 6.7(88) 6.0(1) 4.8(7) 21.0(2) 11.9(26) 80- -89 5.7(46) 4.0(1) 7.2(18) 90- -99 3.7 (14) 5.8(11) 100- -109 3.1 (3) 6.9 12) some indication of the neeti for revising model for- mulations. For example, changes in apparent natu- ral mortality or unexplained changes in fishing mor- tality could be evaluated as the result of losses due to emigration or increases due to immigration. How- ever, stock definition may be further confounded if cod movements have a temporal element and shifts in distribution occur without permanent loss or gain. Acknowledgments The authors express their appreciation to the many individuals who participated in the at-sea tagging operations and in particular to Francois D'Entrement, a fisherman who joined the 1994 survey. We also thank the individuals finding and returning tags for their interest and co-operation. Don Clark and Jay Burnett provided valuable comments on an earlier version of the manuscript. John Neilson provided many help- ful suggestions from a fisheries management perspec- tive. Suggestions from two anonymous reviewers improved the analysis and clarity of the report. Literature cited Clark, D. E. 1996. Assessment of cod in Division 4X in 1996. Cana- dian Department of Fisheries and Oceans (DFO) Atl. Fish. Res. Doc. 96/101, 44 p. Clay, D., W. T. Stobo, B. Beck, and P. C. F. Hurley. 1989. Growth of juvenile pollock (Pollachius iirc/is) along the Atlantic coast of Canada with inferences of inshore-offshore movements. J. Northwest Atl. Fish. Sci. 9:37-43. Gavaris, S. 1988. An adaptive framework for the estimation of popula- tion size. Can. Atl. Fish. Sci. Adv. Com. Res. Doc. 88/29, 12 p. Gavaris, S., L. VanEeckhaute, M-I. Buzeta, and J. Hunt. 1993. Yield projections for the transboundary cod and had- dock resources on Eastern Georges Bank. DFO Atl. Fish. Res. Doc. 93/71, 19 p. GuUand, J. A. 1983. Fish stock assessment: a manual of basic methods. Wiley, Chichester, UK. Halliday, R. G. 1973. Int. Comm. NW. Atl. Fish., Res. Doc. 73/7, serial no. 2909, 19 p. Halliday, R. G and A. T. Pinhorn. 1990. The delimitation of fishing areas in the Northwest Atlantic. .J. Northwest Atl. Fish. Sci. 10:1-50 Halliday, R. G., J. McGlade, R. Mohn, R. N. O'Boyle, and M. Sinclair. 1986. Resource and fishery distributions in the Gulf of Maine area in relation to the Subarea 4/5 boundary. North Atl. Fish. Org. Sci. Council. Studies 10:67-92 Hunt, J. J. 1996. Rates of sexual maturation of Atlantic cod in NAFO Division 5Ze and commercial fishery implications. J. Northwest Atl. Fish. Sci. 18:61-75 Hunt, J. J., and M-I Buzeta. 1989. Status of the Atlantic cod stock on Georges Bank in Unit areas 5Zj and 5Zm. Can. Atl. Fish. Sci. Adv. Com. Res. Doc. 89/47, 26 p. 1996. Biological update of Georges Bank cod in Unit Areas 5Zj,m for 1978-95. DFO Atlantic Fisheries Res. Doc. 96/23, 37 p. Hunt, J. J., and J. D. Neilson. 1993. Is there a separate stock of Atlantic cod in the west- ern side of the Bay of Fundy? North Am. J. Fish. Man- age. 13:421-436 McCracken, F. D. 1956. Cod and haddock tagging off Lockport, Nova Scotia. Fish. Res. Board Can., Prog. Rep. Atl. Coast Stns. no. 64, p. 10-15. McKenzie, R. A. 1956. Atlantic cod tagging off the southern Canadian mainland. Bull. Fish. Res. Board Can. 105, 93 p. NEFSC (Northeast Fisheries Science Center). 1994. Report of the 18th Northeast Regional Stock Assess- ment Workshop. U.S. Dep. Commer., NOAA/NMFS/ NEFSC, Woods Hole, MA. NEFSC Ref Doc. 94-23. 860 Fishery Bulletin 97(4), 1999 Saunders, M. W, G. A. McFarlane, and R. J. Beamish. 1990. Factors that effect the recapture of tagged sablefish off the west coast of Canada. Am. Fish. Soc. Symp. 7:708-713. Scott, W. B., and M. G. Scott. 1988. Atlantic fishes of Canada. Can Bull. Fish. Aquat Sci. 219, 7.31 p. Serchuk, F. M., and S. E. Wigley. 1992. Assessment and management of the Georges Bank cod fishery: an historical review and evaluation. J. North- west Atl. Fish. Sci. 13:2.5-52. Shackell, N. L, W. T. Stobo, K. T. Frank, and D. Brinkman. •1997. Growth of cod ^Gadus morhua ) estimated from mark- recapture programs on the Scotian Shelf and adjacent areas. ICES J. Mar. Sci. 54:383-398. Templeman, W. 1962. Divisions of cod stocks in the northwest Atlantic. In Int. Comm. NW. Atl. Fish., Redbook, Part III, p.79-123. Wise, J. P. 1963. Cod groups in the New England area. U.S. Fish Wild. Serv. Fish. Bull. 63:189-203. Wise, J. P., and A. C. Jensen. 1960. Stocks of the important commercial species offish in the ICNAF Convention area. Int. Comm. NW Atl. Fish. Ann. Meet. Doc. 25, ser. no. 743, p. 1-14. 861 Abstract.— The parameters from von Bertalani'fy's growth equation were es- timated for Mugil cephaluti and M. cureina located in Tamiahua coastal lagoon in Veracruz, Mexico. Also differ- ences in growth and longevity between sexes were obtained. The following were growth parameters for M. cep/wlus: L, = 642.4 mm, W.. = 2.352.12 g, /■- = 0.099, l„ -2.85. (Aogg) = 28.3 years; and for M. cureina: L„ = 461.4 mm. W^ = 844.73 g, /t = 0.14 y t„ = -3.62, (A^ 95) = 18.7 years. Some important differ- ences among growth rates from other areas were found. Significant differ- ences in the growth rate between sexes were observed. The following were growth parameters forM. cephaliis: fe- males: L, = 622.9 mm, k = 0.107, /„= -2.67; Apj,jj = 26.2 years; males; L, = 603.9 mm, A = 0.105,/,, -2.98; A„ 26.5 years; for M. ciirema: females; L„ = 454.6 mm, A' = 0. 135. /„ = -3.94; Ag g^ = 19.2 years; males; L^ = 411.8 mm, /j= 0. 187. /„ = -3.03; A,, 35 = 14.0 years. Growth analysis of striped mullet, Mugil cephalus, and white mullet, M. curema (Pisces: Mugilidae), in the Gulf of Mexico Ana L. Ibanez Aguirre Universidad Autonoma Melropolitana-lztapalapa Ap Postal 55-535, 09340 Mexico, D.F, E mail address anam'xanum uam mx Manuel Gallardo-Cabello Institute de Ciencias del Mar y Limnologia. UNAM Ap. Postal 70-305, 04510 Mexico, D.F. Xavier Chiappa Carrara Unidad de Investigacion en Ecologia Manna, FESZ, UNAM Ap. Postal 9 020, 09230 Mexico, D.F Manuscript accepted 16 December 1998. Fish. Bull. 97;861-872 (1999). The striped mullet, Mugil cephalus (Linnaeus, 1758), has a worldwide distribution, between 42"N and 42°S (de Silva, 1980), whereas the white mullet, Mugil curema (Valen- ciennes, 1836), is basically an American species found from Cape Cod, USA, to Brazil in the Atlantic and from Bahia Magdalena, Mexico, to Chile in the Pacific (Jordan and Everman, 1896). However, Alvarez (1976) has recorded M. curema off the western coast of Africa, In Mexico, 99% of the commercial catch of these mullets takes place in the states of Tamaulipas, mainly in Laguna Madre and in Veracruz, and mainly in the Pueblo Viejo and Tamiahua. The mullet fishery con- stitutes one of the ten most impor- tant fisheries in Mexico as a result of its catch volume, which surpasses 10,000 metric tons (t) annually (Polancoetal., 1987), These species provide both the meat and the roe, locally called "hueva." which commands a gi'eater commei'cial price than the meat. Whereas striped mullet meat has a value of one US$ per kg, the roe is valued at seven US$ per kg. The female gonad has a widespread market because it is eaten region- ally as well as nationally and inter- nationally. In 1995 alone, 43 t of roe with a value of US$ 150,000 (at the present time US$ 800,000) were ex- ported to the United States (Polanco etal,, 1987), Although these species of mullets represent an important source of food in many countries, studies on their ecology and population dy- namics are insufficient. In Mexico some population parameters of M. ceplialus in Tamiahua lagoon, Veracruz, have been analyzed by Marquez (1974) and Garcia ( 1980). Similarly, Diaz and Hernandez (1980) and Romero and Castro (1983) studied M. cephalus in the San Andres lagoon in Tamaulipas and in the Mar Muerto in Chiapas, respectively. Yanez-Ai'ancibia ( 1976) analyzed some aspects of feeding hab- its, growth, and maturity of M. curema in the coastal lagoon system of Guerrero. Comparative studies on M. curema and M. cephalus in Tamiahua lagoon, Veracruz, were carried out by the following authors: Perez-Garcia and Ibanez-Aguirre (1992) and Ibanez-Aguirre and Lleonart ( 1996) on relative growth 862 Fishery Bulletin 97(4), 1999 and comparative morphometries; Ibanez-Aguirre and Gallardo-Cabello (1996a) on total and natural mor- tality and Sanchez-Rueda et al. ( 1997 ) on the analy- sis of sediments in stomach contents. On the inter- national scene, the studies on M. cephalus by de Silva ( 1980 ), Perera and de Silva ( 1978 1, de Silva and Silva (19791 in a coastal lagoon in Sri Lanka, Drake et al. (1984) in the coastal lagoon of San Fernando Cadiz, Cech and Wohlschlag ( 1975) along the coast of Texas, Broadhead (1958) along the coast of Florida, and Thompson (1963) in Australia, are worthy of men- tion, as well as those carried out by Alvarez (1976, 1979 and 1981), Richards and Castagna (1976). and Philhps et al. ( 1987 ) on M. curema . In view of the above studies, the purpose of this study was to carry out an in-depth analysis of the growth characteristics of M. cephalus and M. curema with respect to length, sex, weight, and longevity in Tamiahua lagoon, Veracruz, Mexico. Material and methods Specimens were obtained from the commercial catch landed near Tamiahua lagoon, Veracruz, Mexico ( Fig. 1). The most commonly employed fishing nets were gill nets of 35-mm mesh size (knot-to-knot) for M. 9800' nlet of Tampachiche / Gulf of Mexico 40' ■ Veracruz State 20" Inlet of Corazones Old Inlet 2rOO" cephalus and 30-mm mesh size (knot-to-knot) for M. curema . Sampling was carried out monthly during the first 8 days of each month for a year, from April 1991 to March 1992. Total length (TL) was recorded from 2628 speci- mens of M. cephalus and 3354 of M. curema . For the study of growth, two independent readers examined scales, and otoliths (right sagitta) from 232 speci- mens of M. cephalus (122 females and 110 males) ranging from 200 to 400 mm and 292 specimens of M. curema ( 148 females and 144 males) ranging from 180 to 330 mm. The scales were cleaned, placed be- tween two glass slides, and observed with transmit- ted light. The otoliths were submerged in a petri plate in xylol as a clarifying liquid and observed through a stereoscopic microscope with transmitted light. The analysis of the annual frequency variation of the fast growth rings (opaque) and slow growth rings (hya- line) of the margin of the otoliths showed that each year, one fast growth band and one slow growth band are deposited in the otoliths of boths species of mul- lets (Ibanez-Aguirre and Gallardo-Cabello, 1996b). It was reported earlier that otoliths give better re- sults than do scales for the age determination of both species and make possible a definition of five age groupsfovM. cephalus and sixfor M. curema (Ibanez- Aguirre and Gallardo-Cabello, 1996b). These aver- age lengths were used to obtain the constants for the von Bertalanffy equation (Table 1). Ages "0" and "1" were not collected for M. cephalus, as a result of the mesh size of the nets; instead they were obtained by using the back-calculation method ofLea(1910, in Francis, 1990) and Bagenal andTesch (1978, in Francis, 1990). The constants for the von Bertalanffy equation ( 1938), L^^, k , and /„, were obtained by using the com- bmed method of Ford ( 1933 ) and Walford ( 1946), and the methods of Gulland (1964), Tomlinson and Abramson (1961), Allen (1966), Beverton^, Pragei ( 1987 ), and Bayley ( 1977 ). Growth was also measured for each sex of both species. Growth curves for both species were obtained from the constants of the equa- tion calculated by the above mentioned methods. The sum of the squared differences (Ze,'^) was used to compare the differences between calculated and ob- sei"ved values. Hotelling's T' test (Bernard, 1981) was used to compare growth curves of the two sexes of both spe- cies. This test assumes that estimations ofL. , k, and /„ for both groups were obtained from two normal Figure 1 Map of the Tamiahua Lagoon, Veracruz, Mexico. ' Beverton, R. J. H. 19.54. Notes on the use of theoretical mod- els in the study of the dynamics of exploited fish populations. U.S. Fish Wildl. Serv., Fisheries Laboratory, Beaufort. Misc. ("ontrib. Rep., 181 p. Ibahez Aguirre et al,: Growth analysis of Mugil cephalus and M. curema 863 Table 1 Mean length (mm) employed for obtaining the growth parameters. Mugil cephalus Mugil curema Females Males Species Calculated Females Males Species Calculated Age (±SD) (±SD) (±SD) n values (±SD) (±SD) (±SD) (±SD) n values (±SD) "0" 156'(± 91 160'(± 13) 158" (± 11) 0 158.34 <± 9) 187 (± 10) 176(± 11) 183 (±4) 4 184.20 (±9) "1" 203'(± 7) 207' (±9) 204'"(± 7) 0 204.09 (±9) 224 (±8) 221 (±9) 223 (± 7) 60 220.56 (± 11) "2" 246 (±5) 247 (± 19) 246 (± 14) 45 245.53 (±13) 251 (±6) 250 (± 8) 252 (±9) 71 252.15 (±6) "3" 283 (± 3) 281(± 10) 282 (± 11) 59 283.04 (± 11) 276 (± 7) 276 (± 7) 278 (±8) 66 279.60 (±4) "4" 319 (± 7) 315 (± lOl 317 (± 9) 44 317.01(±11) 299 (± 6) 300 (±4) 303 (± 7) 54 303.44 (± 3) "5" 350 (± 9) 344 (±8) 349 (± 12) 51 347.77 (± 8) 320 (± 4) 321 (±4) 325 (± 8) 37 324.16 (±4) "6" 377 (±8) 369 (±6) 375 (± 12) 33 375.62 (± 11) — — — ' Average lengths obtained by back calculation. Table 2 Estimates of the constants of the von Bertalanffy equation for M. ce phalus. according to the different methods used. Method Species Females Males LJ.mm, k 'o SD2 LJ.mm) k 'o SD2 Ljmm) k to SD2 Walford-Gulland 637.97 0.0998 -2.8838 4.049 627.54 0.1052 -2.7342 2.186 609.05 0.1028 -3.0651 3.339 Beverton regression 637.97 0.1000 -2.8723 3.827 627.54 0.1053 -2.7281 2.110 609.05 0.1029 -3.0575 3.237 Tomlinson and Abramson 640.64 0.0998 -2.8418 3.211 627.24 0.1054 -2.7280 2.115 604.93 0.1043 -3.0376 3.341 Beverton regression 640.64 0.0998 -2.8400 3.211 627.24 0.1059 -2.6958 1.793 604.93 0.1050 -2.9860 2.725 Allen 642.00 0.0993 -2.8483 3.421 622.87 0.1074 -2.6709 1.794 603.52 0.1054 -2.9772 2.919 Beverton regression 642.00 0.0994 -2.8476 3.210 622.87 0.1074 -2.6697 1.783 603.52 0 1055 -2.9765 2.723 Prager 642.40 0.0993 -2.8480 3.210 622.90 0.1074 -2.6690 1.790 603.90 0.1054 -2.9770 2.722 Beverton regression 642.40 0.0993 -2.8499 3.209' 622.90 0.1074 -2.6699 1.782' 603.90 0.1054 -2.9791 2.722' Bayley 640.00 0.1003 -2.8210 3.408 615.89 0.1094 -2.6620 2.382 608.34 0.1055 -2.8988 8.592 ' The best fit. distributions of joint probability, with three variables and one common variance. The complete and eviscerated weights of 473 and 329 specimens of M. cephalus and M. curema , re- spectively, were recorded for the study of growth by weight. Weight data were recorded for each 20-mm length interval and the average weight for each size class was calculated. The function W= aL * was used to obtain the weight- length relationship. Data for growth by length and the weight-length relationship were used to obtain the weight for each age. Growth by weight was ob- tained by substituting L, and L^ by W, and W„, re- spectively, obtained from the weight-length relation- ship in the von Bertalanffy equation. Taylor's equa- tion (1958, 1960) was used to calculate age limit or longevity (95% oiL^). Results Growth in length Mugil cephalus The values for L ., and k were very similar ( Table 2 ). The method that provided the great- est differences was that of Ford (1933) and Walford (1946). This similarity was due to the obtention of these constants; the t^ value determined by Gulland 864 Fishery Bulletin 97(4), 1999 (1964) was also employed. Gulland recommends us- ing just the age groups that are best represented and thus avoids errors due to a low representation of some poorly sampled age groups. Some calculations were repeated and the following tg values were obtained for the 3-5 years age groups: -2.74 (females), -3.07 (males), -2.89 (both species, sexes combined); and for the 3-6 age groups: -2.67 (females), -2.98 (males), -2.85 (both species, sexes combined). The best fit was found with the t^ result calculated for 3-6 years age groups. Thus, these values were used to obtain the average ^q (Table 2). Use of the Beverton^ equation improved the calculated values in comparison with the values observed by Ford (1933) Walford (1946), Tomlinson and Abramson ( 1961 ), Allen ( 1966 ), Bayley ( 1977 ), and Prager ( 1987 ) methods. The calculated curve that best fitted values ob- served through otoliths corresponded to the para- meters that were calculated with the Prager (1987) method and fitted with the Beverton^ equation (Table 2). The calculated values of the lengths for different ages as well as their standard deviation (SD) were obtained by using these parameters for the two spe- cies (Table 1) which are consistent and, in general, show improved calculated values. Figure 2 presents the theoretical growth curve for M. cephalus ages 0-6 years. During the first two years of life, striped mullet grew rapidly in length, with average increases of 45.8 mm during the first year and 41.4 mm during the second. From the third year on, growth decreased to annual increases in total length of 37.5 mm. Increases 400 Observed values » Calculated M. cephalus values ^- 350 .■'■^ ? 300 E t 250 & y^-'" M. ciirema 200 150 0 1 2 3 4 5 6 Age (-years) Figure 2 Theoretica growth curve for M. cephalus and M. vurema. between the third and fifth years varied from 34.0 to 30.8 mm. Between the fifth and sixth years the in- crease was even smaller, with an average increase of 27.8 mm. In general, growth was high during the first two years of life and then decreased. This de- crease is probably related to the time of first sexual maturity, which for this species occurs from 280 to 299 mm TL (males and females, respectively), which corresponds to an age of 3 years in both cases (Ibahez- Aguirre and Gallardo-Cabello, 1996b). L^and k have a negative correlation, whereas L,., is high, the growth rate is low (Table 2). Mugil curema As can be seen in Table 3, L„ and k values obtained with the Ford (1933) and Walford ( 1946) method show the greatest difference with re- spect to the values calculated by the Tomlinson and Abramson (1961), Allen (1966), Prager (1987), Beverton,' and Bayley (1977) methods. As with M. cephalus , new calculations for the t^ value were made by using the Gulland ( 1964 ) method for the 1-4, 3-5, and 2-4 years age groups; the fol- lowing t^ values were obtained: for the 1-4 years age groups -3.94 (females), -3.03 (males), -3.62 (both species, sexes combined), for the 3-5 years age groups: -3.73 (females), -2.77 (males), -3.41 (both species, sexes combined) and for the 2-4 years age groups -3.72 (females), -2.75 (males), -3.39 (both species, sexes combined). The von Bertalanffy curves showed the best fit for the calculated values of the 1-4 yr age groups. These values were used to obtain average Iq. The Beverton' equation with the above methods improved calcu- lated values only in the case of the Ford (1933) and Walford (1946) method. The calculated curve that best fitted observed values through otoliths corre- sponded to the parameters calculated with the Prager ( 1987 ) method, as can be seen in Table 3. Using these parameters, we calculated values for lengths at different ages, as well as their SD, which, as in the case of M. cephalus, are consistent and, in general, show improved calculated values (Table 1). Figure 2 presents the theoretical curve for growth in length of M. curema , for the ages of 0-5 years. A high increase in length was recorded during the first year, after which growth decreased markedly. Size increased by 27.5 mm TL between the second and third years, 23.8 mm between the third and fourth years, and 20.7 mm between the fourth and fifth years. The decrease in growth from the first year on is related to the first sexual maturity, which in this species occurs in small sizes from 181 to 208 mm TL for males and females, respectively, at ages "0" and "1" (Ibaiiez-Aguirre and Gallardo-Cabello, 1996b). Ibaiiez Aguirre et al,; Growth analysis of Mugil cephalus and M. curema 865 Table 3 Estimates of the constants of the von Bert alanffy equation for M. curema. according to the different methods employed. Method Species Females Males LJmm k ■»/■.■ jr 0.6 op '* ■v*' : '3j t3 ■ •• ^t^^" 2 0.4. • .irf*»>-- ' ■■ 4J jffiP^ ■ ^■f^^ "> 0.2- _adM^^. tu .^^^Ir^ . .,>f**^ 0- 240 280 320 360 400 440 Length (mm) Figure 3 Relation between weight and length of M. ccphaliis: (A) com- plete we ght; (B) evi.scerate(i weight. Finally, it is important to take into account that differences between growth rates are important even in areas that are very close; these differences could be explained by the different methods applied for age determination (Oren, 1981); however, these differ- ences could be also explained by the world-wide dis- tribution of this species and its different survival strategies. On the other hand, the differences be- tween growth rates can also occur because of com- mercial exploitation, for when fishing is very intense, the commercial size offish decreases and the varia- tions of the k coefficient increase. The values of the parameters of the von Bertalanffy growth equation for M. curema in different areas are shown in Table 7. In our paper, the value of/? calcu- lated for M. curema shows a higher position in rela- tion to the value proposed by Alvarez ( 1979) and the value of L^_, is higher in Cuba. On the other hand, the values of A' obtained by Richards and Castagna ( 1976) in Virginia and by Phillips et al. (1987) in theNicoya Gulf of Costa Rica show a higher position in relation to the k value obtained in our paper. For the relation of growth between sexes, some authors have shown that there are no differences between sexes: for M. cephalus, Dannevig (1902); Kesteven (1942); Thomson (1951); Morovic (1957); Erman ( 1959); Thakur ( 1967); Cech and Wohlschlage (1975); Grant and Spain (1975); for M. curema, Alvarez ( 1979) and Angell (1973). On the other hand, Ezzat (1965), Brulhet (1974; 1975), and Farrugio (1975) have stated that there are differences in the growth between sexes. However, these latter authors did not infer whether these differences in the growth between sexes are significant or not from a statisti- cal point of view. For this reason, a statistical test to compare the growth curves between sexes was 868 Fishery Bulletin 97(4), 1999 400i A W= 3.79 X 10-5 (TL)-'^5 ■ n=465 • .■• / ? 300- 1 r=0.99 ■ S>' ■ Afifc*' ■ u JK4 ^ 200- rtlK^ o. E o U ^■' 100 ■ Q 160 200 240 280 320 360 380 T B " W= 1.06 X 10-5 (TL)2-94 ■ ■ _ 320- n=330 ~29 r=0.87 ■■ ■ 'i' 260- flj ^"" ■■ rfW ■ S ^'■' ■ T3 1 200- u mT^ o J6«^ iS 140- - ■ -aK*^- ■ AftM^' 80- .Jl^^ 4 160 200 240 280 320 360 Length (mm) Figure 4 Relation between weight and length of M. ctirema: (A) com- plete weight; (B) eviscerated weight. applied in our study (Tables 4 and 5); the results show that there are significant differences in growth be- tween males and females for M. cephalus and M. curema . Oren ( 1981) mentioned: "sometimes the fe- males grow slightly faster (Thomson, 1951; Hickling, 1970); Cech and Wohlschlag, 1975), live longer than the males (Thomson, 1951) or at least are predomi- nant among older fish (Hickling, 1970)." In general, the values of the relationship between length and weight obtained in our study are very similar to those expressed by other authors for coastal lagoons and marine areas. The results obtained for M. cephalus are similar to those shown by Kesteven (1942), Morovic (1954), Marquez (1974), Serbetis (1939), and Ezzat (1965). In the same way, the re- sults obtained for M. curema are similar to those shown by Angell (1973) and Richards and Castagna (1976). The longevity values for M. cephalus in different localities are given in Table 6. The highest values, 57.6 and 49.9 years, were found in the marine zones by Ilin ( 1949) in the Black Sea and Kesteven (1942) in Australia, respectively. The lowest values were obtained by Broadhead (1958), 3.7 for males and 4.5 years for females of this species in the marine zones, and by Heldt (1948), 4.6 years in coastal lagoons. The values of longevity for M. curema in different areas are shown in the Table 7. The highest value was obtained by Alvarez (1979), 30 years in Cuba, and the lowest by Richards and Castagna ( 1976), 3.8 years in Virginia. The longevity values obtained in the study, 28.3 and 18.7 years for M. cep/!o/!vs and M. curema. re- spectively, showed an intermediate position in rela- tion to the values found by other authors. In all cases, as Taylor (1958) has shown, the longevity and Ibanez Aguirre et al,: Growth analysis o\ Mugil cephalus and M. curema 869 A ° Calculated values • Observed values 600 oo Wt= 2,352.12 (l-e-0 1'(t+2,85))2.80 ..a £ 500 , DO ^x" S 400 y^ ■E. 300 ^^ E ^^ o O 200- j^'" 100- -^ 12 3 4 5 6 7 B 500i QO Wt= 1,986.27 (l-e-0"(t+2.85))2.86 y •S, 400 y' u X s ^ ■a 300- y* B ^/ a 200- ^ > ■r- UJ 100- ,i 12 3 4 5 6 7 Age (years) Figure 5 Theoretical growth cune of A/, cepbalus: (A) com plete weight; (B) eviscerated weight. 400 00 r 300 op E 200 1 o- ,^ 100 t Calculated values < Observed values W,= 844.73 (l-e-'>'4(t+3.62))2.75 M 300 100 B Wt= 736.89 (l-e-'"4(t+3.62))2.94 12 3 4 5 6 7 Age (years) Figure 6 Theoretical growth curve of M. curema: (A) complete weight; (B) eviscerated weight. Table 6 Growth parameters of M . cepi ali/s for the Gulf of M exico and other localities. TL = total length. F = fem ale, M = males. Sp. = species. This table was modified from Tables 5.4 and .5.5 in Oren (1981). Authority Locality Method Length Sex L, k 'o A I ^0 95 Gulf of Mexico Coastal lagoons This study Tamiahua, Mexico Otoliths TL F 622.9 0.11 -2.670 26.2 Otoliths TL M 603.9 0.11 -2.979 26.5 Otoliths TL Sp. 642.4 0.10 -2.850 28.3 Marquez. 1974 Tamiahua. Mexico Scales TL Sp. 510.0 0.34 -0.114 8.8 Diaz and Hernandez, 1980 Tamaulipas, Mexico Scales TL Sp. 588.0 0.19 -0.213 15.4 Marine zones Cech and Wohlschlag, 1975 Texas, USA Scales TL F 407.0 0.32 -0.710 9.4 Scales TL Sp. 450.0 0.24 -0.900 12.5 Broadhead. 1958 N & NW Florida USA Scales and tag TL F 374.0 0.82 -0.160 3.7 Scales and tag TL M 379.0 0.66 -0.036 4.5 Other localities Marine zones Ilin, 1949 Black Sea Scales TL Sp. 1089.0 0.05 -1.620 57.6 Kesteven. 1942 Australia Scales TL Sp. 1729.0 0.06 -0.510 49.9 Thompson, 1951 West Australia Scales TL Sp. 609.0 0.30 -0.143 10.0 Thompson, 196.3 Australia Scales TL Sp. 727.0 0.23 0.006 con 13.1 tinui'd 870 Fishery Bulletin 97(4), 1999 Table 6 (continued) Authority Locality Method Length Sex L_ k to A ' Coastal lagoons Romero and Castro, 1983 Chiapas, Mexico Scales TL- Sp 458.5 0.21 -1.770 14.4 Ezzat, 1964 France Otoliths TL Sp 417.7 0.47 -0.169 6.4 Serbetis, 1939 Rome, Italy Scales TL Sp 563.0 0.56 0.083 5.3 Morovic, 1954 Venice, Italy Scales TL Sp 611.0 0.21 -0.465 14.3 Alessio, 1976 Orbetello, Italy Scales TL Sp 615.0 0.40 -0.044 7.5 Morovic, 19.57 Vransko, Yugoslavia Scales TL Sp 590.0 0.23 -0.083 12.8 Heldt, 1948 Tunisia Scales TL Sp 620.4 0.65 -0.048 4.6 Farrugio, 1975 Tunisia Scales TL Sp 693.0 0.19 -0.630 15.8 ' These values of longevity were obtained in our study by th authors mentioned in this table. ^ For the conversion from standard length to TL, the equation ; application of the given by Thompson Taylor method 1 19581 to the grow et al (19911 was used th para meters given by the Table 7 Growth parameters of M. curema for other localities. Authority Locality Method Length Sex L„ k 82 cm) length was highly significant (38% vs. 71%; x^=4-36, df=l, P<0.025). Figure 4 shows observed postbiting behavior tran- sitions, from an initial complete bite by an individual fish through the fourth behavioral transition. Of 50 1=' Rushing 47 Spitting 2 Hooking 1 Biting Spitting 23 Hooking 20 Departure 1 3rd 1 Looping [6 Rushing 1 Departure 17 41H Figure 4 Behavior tree describing behavior sequences following an ini- tial complete bite. Numbers at left of rows are the position of behaviors in sequences following initial observation. Numbers in boxes are the total frequencies of each behavior in a row position. Thickness of arrows represents relative numbers of observations for each behavior transition. Because some be- haviors were compromised, not all behaviors have transitions from one row position to the next. Table 4 Halibut attack rate 'r offish biting) and hooking success i^f of bites resulting in hooked fish) by 5-cm group. Size Number class Number of % Number % (cm) observed bites bites hooked hooked 62-67 6 3 50 0 0 68-72 11 5 45 1 20 73-77 17 9' 53 3 38 78-82 11 8 73 5 63 83-87 12 5 42 3 60 88-92 18 6 33 4 67 93-97 12 6 50 5 83 Total 87 42 21 ' One bi e in the 73-7 7 cm size c ass was compromised. Percent- age hooked is determined from the eight uncompromised bites. fish tracked on this chart, 20 ended up departing, 22 ended up hooked, and 3 were still interacting with hooks at the end of the fourth behavioral transition. Of the three continuing interactions, one fish lay near bait and two fish rebit baited hooks. Fifty percent of 880 Fishery Bulletin 97(4), 1999 spits occurred in less than 5 seconds, and 95% oc- curred in less than 25 seconds. Once hooked, and after initial rushing, the halibut lay on the bottom, resting, then went into a pattern of rushing, resting, rushing, etc. There appeared to be a pattern of de- creasing duration of rushing following subsequent rest periods. The average times for resting and rush- ing after the initial hooking were 3 min 49 s, and 31 s, respectively (Table 2). Hooking success for other species The range of approach and interaction behaviors for species other than halibut was not documented be- yond noting bitings and subsequent hooking. Other species that bit at hooks included canary rockfish iSebastes pinniger), yelloweye rockfish iS. ruberrim us ), quillback rockfish iS. maliger), raffish (Hydrolagus colliei), and lingcod (Ophiodon elongatus). In general, hooking success for the major rockfish species was around 5-6% (Table 5). These fish were comparable in average size (weight) to those caught as inciden- tal catch in the directed halibut fishery. Discussion The feeding behavior of halibut may be classified into three phases; arousal, search, and bait attack and food ingestion. Arousal In most cases, arousal to the presence of food occurs at a distance, and bait odor carried by bottom cur- rents attracts fish from well beyond the limited view- ing distance of the present experiment. Although undoubtedly involving detection of an odor plume (Atema, 1980), the present experiment did not investi- gate the initial arousal phase of feeding behavior Table 5 Hooking success for species other than hahbut. Average Number Aieight (kg) of Number ^f of hooked Species bites hooked hooked fish Canary rockfish 339 17 5 4.0 Yelloweye rockfish 237 20, 8 3.8 Quillback rockfish 228 9 4 1.9 Ratfish 36 1 3 N/A Lingcod 23 9 39 11.1 However, a number of observations in relation to food location and uptake were made. Search It is clear that halibut use orientation to bottom cur- ' rent to locate food, most approaching upstream when a bottom current was present. This rheotactic orien- tation is used by many fish for detection and loca- tion of prey (Atema, 1980; L0kkeborg et al., 1989; Lokkeborg, 1998). Once fish are aroused to the pres- ence of a prey item by scent, they orient into a cur- rent to locate the food. The role of vision in food location by Pacific hali- but was less clear. Behaviors directed towards the gear were observed in some cases immediately after the gear reached the seafloor. These were likely re- sponses to visual clues. However, the greatest num- ber of appearances appeared to be motivated by scent carried by the current. Although earlier studies have shown that halibut certainly prey on a number of pelagic or semipelagic species where vision must play an important role in prey recognition and capture (Best and St-Pierre, 1986), it probably played a lim- ited role in behavior toward the model setline. Prebiting behavior and attack rate Fish vary widely in their reliance on sight, smell, and touch in deciding whether to accept or reject food once it has been located ( Lokkeborg, 1994 ). Both loop- ing and lying behaviors were common in the observed halibut, and in only a few cases was a complete bite initiated without some preliminary bait interaction such as looping or lying. Looping behavior could be a test of an odor plume by the fish, assuring that the fish is in an area of high-scent concentration. Many halibut lay near the bait prior to initiating a bait attack, most either directly downstream or to one side of the bait in relation to the current. This behav- ioral response may reinforce olfactory clues that led the fish to the bait, or possibly a restrained response to the baited hook as a novel prey item (Lokkeborg, 1990). Rejection of the bait at this point was often followed by looping, behavior that often led to further bait interaction. The role of mechanoreception was not demonstrated. In only a few instances did halibut exhibit incomplete bites, and these generally did not result in the fish then leaving the area of the gear. Most bites were associated with extremely active be- havior, e.g. a fish beginning to rush simultaneously with biting the bait. Only one bite that resulted in hooking was not followed by a rush. In the simplest interpretation, the attack rate for halibut was 43%. Fish length was not a significant Kaimmer: Hooking behavior of Hippoglossus stenolepis factor in attack rate, although approach direction was highly significant; fish that approached the bait up- stream bit almost twice as often as fish that ap- proached from the side or downstream, and four times as often as fish that approached during slack current. It is likely that fish approaching upstream were following a scent trail from the baits and had a higher motivation to bite than fish that approached from other directions. It is interesting to note that although halibut ap- peared throughout the gear deployment, consistent with the idea of being aroused by and following a scent plume from different distances, once a fish ar- rived at the area of the baited hooks, bait attacks occurred quickly. Most attacks occurred within the first minute after a halibut appeared, and less than five percent of the bait attacks occurred more than one and one-half minutes after the fish appeared. Postbiting behavior and hooking success In almost all cases, complete bites were followed by rushing behavior, which resulted in about equal num- bers of hooked fish and fishes that spat the hook and then departed the observation area. Hooked fish struggled violently for a short period, then went into a series of resting and rushing behaviors which con- tinued through the observation period, the duration of rushes becoming gradually shorter. Size selectivity by hook-and-line gear has a num- ber of components, including encounter rate, attack rate, and hooking success. A higher encounter rate has been shown for larger fish, which may have greater foraging ranges and therefore a higher prob- ability of encountering baited gear ( Lokkeborg and Bjordal, 1992; Engas and L0kkeborg, 1994; Bjordal and Lokkeborg, 1996). It is possible that the present experimental arrangement was skewed in this way toward catching larger halibut. Attack rate has been related to bait size for some species; larger fish show a preference, and therefore a higher attack rate, for larger baits, and smaller fish showing a preference for smaller baits (Bjordal and Lokkeborg, 1996). No difference in attack rate by fish size was seen in the present study. Hooking success was found to be strongly depen- dent on fish length, ranging from zero for fish less than 62 cm, to 839. for fish 93-97 cm (Fig. 5). This finding is consistent with and can explain much of the selectivity estimated from representative IPHC commercial and setline survey data (Clark^). A circle 0% > 65 70 75 80 85 90 95 Fork Length (5-i:iii interval) • - Observed hooking success — Selectivity estimated from survey Figure 5 Hooking success from direct observation and se- lectivity estimated from IPHC survey data by .5-cm length groups. ' Clark, W. 1997. Coastwide distribution of exploitable bio- mass according to 1997 setline surveys. Int. Pac. Halibut • Comm. Rep. of Assessment and Research Activities, p. 161- 202. International Pacific Halibut Commission. P.O. Box 9.5009, Seattle, WA 9810.5-2009. Unpubl. manuscript. hook is designed so that it is pulled to the corner of the mouth during rushing, with the point of the barb in line with the pull of the gangion (Bjordal and Lokkeborg, 1996). Hooking results from the orienta- tion of the hook during the rush and the penetration of the barb caused by the pull on the gangion during that rush, the point of the hook circling the jawbone and exiting through the cheek (Johannes, 1981). Ninety to ninety-five percent of halibut caught on circle hooks are hooked in this manner (Kaimmer andTrumble, 1997). Hooking success has been related to hook size only when the range of hook sizes is very great, on the order of 2009^ or more (L0kkeborg and Bjordal, 1992) and has been explained in terms of larger fish being able to exert a stronger pull on the gangion during rushing. This stronger pull generates the greater force necessary to pull the point of a larger hook fully into the tissue of the mouth cavity, resulting in a higher rate of hooking success. A greater hooking success as the result of a larger fish exerting a stron- ger pull may be countered by weaker tissue in the mouth of smaller fish, requiring less force for the hook to penetrate (Bjordal and Lokkeborg, 1996). The hooking success table (Table 4) in this study was con- structed for fish 62-97 cm in length, the largest just 1.5 times the length of the smallest, and showed dra- matic differences in hooking success over length changes as small as 10-20 cm. Some mechanism be- yond pull strength was probably responsible for these differences. In an earlier study where the same hooks 882 Fishery Bulletin 97(4), 1999 (Kaimmer, 1994) were used, halibut less than 82 cm in length were hooked more often in locations other than the jaw than were larger halibut. Although this difference was small, it suggests a functional rela- tion between the mechanical operation of the hook and the size of the fish's mouth in relation to overall hook and bait dimensions. A higher rate of hooking success for smaller halibut might have been seen if smaller bait had been used. This was not tested in the present experiment because bait size was held constant to the standard used in IPHC setline sur- veys and representative of the commercial halibut fishery. It is clear that an increased hooking success for larger fish has a dramatic effect on the size selection of longline gear. Much of the length-based selectiv- ity assumed for halibut setlines can be explained by the differences in hooking success by fish length dem- onstrated by this study. The selection curve gener- ated from the present observations is the first to be determined for Pacific halibut through direct ob- servation of hook attacks. The large number of bait attacks by species other than halibut compensated for the very low hooking success for these species. In fact, we caught more of these species than halibut during the experiment. Although their presence or interaction with baits short of hooking did not seem to affect the attack rate of halibut that were present, the removal of available baits by interspecies competition should be included in any model of hook-and-line CPUE. Acknowledgments The author expresses his thanks to S. Lokkeborg for comments on an early draft of this paper, as well as three anonymous reviewers who gave many useful suggestions. I would also like to thank Craig Rose and Scott Mclntire, National Marine Fisheries Ser- vice, Seattle, for loan of the camera gear and for in- valuable assistance in training me in its use, and most especially for their long-distance troubleshoot- ing during the project. Conclusions We consider the results of this study to be qualita- tive, not quantitative. Their application to commer- cial or experimental longline sets or CPUE indices should take into account the limited nature of the experiment, both in terms of number of hooks fished and in terms of lack of seasonal or wide areal varia- tion in the observations. Although it is likely that the results are generally applicable to halibut, they could vary significantly for fish with different feed- ing histories in a different life stage. Attack rate par- ticularly could be susceptible to the condition offish in relation to recent feeding or spawning activity. The primary objective of this project was to determine estimates for the attack rate and hooking success of Pacific halibut on gear typical of those used in the commercial fishery. Results show that although these parameters may be estimated, their values are in- fluenced by a number of factors, most notably by the presence of bottom current and direction of approach in relation to that current (for attack rate) and fish length (for hooking success ). The behavioral sequences observed for Pacific halibut were fairly limited and in- cluded searching into a current for food, looping and lying around or near bait items, and a vigorous biting- rushing sequence that often resulted in hooking. The relative absence of physical contact with the bait prior to biting (such as tasting) might indicate that texture is less important for food selection by halibut. The role of vision in halibut feeding was not tested. Literature cited Atema, J. 1980. Chemical senses, chemical signals and feeding behav- ior m fishes. In J. E. Bardach, J. J. Magnuson, R. C. May, and J. M. Reinhart (eds.l. Fish behavior and its use in the capture and culture of fishes, p. .57-101. Int. Center for Liv- ing Aquatic Resources Management, Manila, Philippines. Best, E., and G. St-Pierre. 1986. Pacific halibut as predator and prey. Int. Pac. Hali- but Comm. Tech. Rep. 21, 21 p. Bjordal, A, and S. Lekkeborg. 1996. Longlining. Fishing News Books, Oxford. 156 p. Deriso, R. B., and A. Parma. 1987. On the odds ofcatching fish with angling gear. Trans. Am. Fish. .Soc. 116:244-256. Engas, A., and S. Lokkeborg. 1994. Abundance estimation using bottom gillnet and longline — the role offish behavior /;; A. Femb and S. Olsen (eds.l. Marine fish behavior in capture and abundance esti- mation, p. 9-27. Fishing News Books, Oxford. 221 p. Ferno, A., P. Solemdal, and S. Tilseth. 1986. Field studies on the behavior of whiting [Gadus merlangus L.) towards baited hooks. Fiskdir. Skr. Ser. Havunders. 18:83-95. Hoag, S. H., R. B. Deriso, and G. St-Pierre. 1984. Recent changes in halibut CPUE: studies on area dif- ferences in setline catchability. Int. Pac. Halibut Comm. Sci. Rep. 71, Seattle, WA, 44 p. Huse, I., and A. Ferno. 1990. Fish behavior studies as an aid to improved longline hook design. Fish. Res. 9:287-297. Johannes, R. E. 1981. Words of the lagoon — fishing and marine lore in the district of Micronesia. Univ. California Press, Los Ange- les, CA. 207 p. Kaimmer: Hooking behavior of Hippoglossus stenolepis 883 Kaimmer, S. M. 1994. Halibut injury and mortality associated with manual and automated removal from setline hooks. Fish. Res. 20:165-179. Kaimmer, S. M., and G. St-Pierre. 1993. Further studies of area differences in setline catcha- bility of Pacific halibut. Int. Pac. Halibut Comm. Sci. Rep. 77, Seattle, WA, 59 p. Kaimmer, S. M., and R. J. Trumble. 1997. Survival of Pacific halibut released from longlines. In Fisheries bycatch — consequences and management; proceed- ings of the symposium on the consequences and manage- ment of fisheries bycatch, 27-28 August 1996, Dearborn, MI. Alaska Sea Grant College Program Rep. 97-02, Univ. of Alaska, Fairbanks. AK, p. 101-106. Lekkeborg, S. 1990. Reduced catch of under-sized cod (Gadus morhua) in longlining by using artificial bait. Can. .J. Fish. Aquat. Sci. 47:1,112-1,115. 1994. Fish behavior and longlining. In A. Ferno and S. Olsen (eds.). Marine fish behavior in capture and abun- dance estimation, p. 9-27. Fishing News Books, Oxford, 1998. Feeding behavior of cod iGadus morhua): activity rhythm and chemically mediated food search. Animal Behavior 56(2):377-378. Lokkeborg, S., and A. Bjordal. 1992. Species and size selectivity in longline fishing; a review. Fish. Res. 13; 311-322. Lokkeborg, S., A. Bjordal, and A. Ferno. 1989. Responses of cod i-Gadus morhua) and haddock {Mclanogrammus aeglefinus) to baited hooks in the natu- ral environment. Can. J. Fish. Aquat. Sci. 46:1,478-1,483. 1993. The reliability and value of studies of fish behavior in long-line gear research. ICES Mar. Sci. Symp. 196: 41-46. Myhre, R. J. 1969. Gear selection and Pacific halibut. Int. Pac. Hali- but Comm. Rep. 51, Seattle, WA, 35 p. Slater, P. J. B. 1973. Describing sequences of behaviour. In P. Bateson and P. Klopfer (eds.). Perspectives in ethology, p. 131- 153. Plenum Press, New York, NY. Skud, B. E. 1978. Factors affecting longline catch and effort: III. Bait loss and competition. Int. Pac. Halibut Comm. Sci. Rep. 64, Seattle, WA, 66 p. 884 Abstract.— Genetic population struc- ture in Atlantic croaker (Micropogonias undulatus Linnaeus) was examined by using the polymerase chain reaction (PCR) and restriction fragment length polymorphism (RFLP) analysis of mi- tochondrial DNA (mtDNA). Juvenile croaker from three U.S. Atlantic locali- ties (Delaware. North Carolina, and Florida) and one Gulf of Mexico local- ity (Louisiana) were screened to docu- ment the magnitude and spatial distri- bution of mtDNA variation in M. undulatus; to evaluate the integrity of Cape Hatteras, North Carolina, as a genetic stock boundary; and to estimate levels of gene flow among Atlantic lo- calities to provide an improved basis for future decisions regarding coastwide management of this fishery resource. RFLP analysis of the ATPase 6 and D-loop mtDNA regions revealed a total of 15 composite haplotypes in 93 indi- viduals. Monte Carlo simulations re- vealed no geographic heterogeneity in mtDNA haplotype frequencies among Atlantic localities and no evidence that juveniles collected north and south of Cape Hatteras originated from sepa- rate gene pools (net sequence diver- gence=-0.002'7f ). There was significant heterogeneity between Atlantic and Gulf of Mexico samples, suggesting re- stricted gene flow between these two re- gions. Analysis of molecular variance also indicated regional (Atlantic versus Gulf) population structure, but pro- vided no evidence that Cape Hatteras represents a genetic stock boundary. AMOVA indicated relatively high gene flow iN^m = 12-23 effective female mi- grants per generation) among Atlantic localities. These findings are consistent with 1) a single genetic stock of M. undulatus on the Atlantic coast and 2) separate, weakly differentiated stocks in the Atlantic and Gulf of Mexico. Mitochondrial DNA analysis of population structure in the Atlantic croaker, Micropogonias undulatus (Perciformes: Sciaenidae) Thomas E. Lankford Jr. Timothy E. Targett Patrick M. Gaffney Graduate College of Marine Studies University of Delaware 700 Pilottown Road Lewes, Delaware 19958 Present address (for T E Lankford Jr) E mail address (for T E Lankford Jr) Marine Sciences Research Center State University of New York Stony Brook, New York 11794-5000 ankford g ccmail sunysb edu Manuscript accepted 24 March 1999. Fish. Bull. 97:884-890 (1999), Atlantic croaker (Micropogonias undulatus Linnaeus) is an impor- tant commercial and recreational fishery species along the U,S. Atlan- tic and Gulf of Mexico coasts (Chao and Musick, 1977; MercerM, On the Atlantic coast, M. undulatus is com- mon in estuarine and coastal waters from Indian River, Florida, to Chesapeake Bay (Nelson et al., 1991; Stone et al., 1994). Current fishery management practices for M. undulatus assume a single or "unit" stock for the entire Atlantic coast (Kline and Speir, 1993) despite indications that two stocks may ex- ist. The two-stock hypothesis is based on variation in life history traits and population dynamics of Atlantic croaker occurring north and south of Cape Hatteras, North Carolina. White and Chittenden (1977) reported lower mortality rates, greater longevity, and larger maximum size for M. undulatus north of Cape Hatteras. Ross ( 1988) concluded that two groups with con- trasting life histories may overlap in the vicinity of Cape Hatteras, with the northern, mainly offshore group exhibiting greater size-at- age, greater longevity, lower annual mortality, and delayed maturation compared with southern individu- als. Geographic variation in biologi- cal characteristics of northern and southern M. undulatus implies that these groups may respond indepen- dently to exploitation and may re- quire management as separate stocks; however, the basis for these differences (i.e. genetic or ecopheno- typic) is unclear. Identification of Atlantic croaker stock structure was listed among the priority research needs by the 1981 Sciaenid Assessment Work- shop sponsored by the Atlantic States Marine Fisheries Commis- sion (ASMFC) and National Marine Fisheries Service ( Wilk and Austin, 1981). The Atlantic Croaker Fish- ery Management Plan, prepared under the ASMFC Interstate Fish- ery Management Program, also rec- ommended research to identify croaker stocks along the Atlantic coast (MercerM. In the present study, polymerase chain reaction (PCR) and restriction fragment length polymorphism analysis ' Mercer, L. P. 1987. Fishery manage- ment plan for Atlantic croaker. Fisheries Management Rep. 10, Atlantic States Ma- rine Fisheries Commission, Washington, DC, 90 p. Lankford et al : Population structure in Micropogonias undulatus 885 Table 1 Sampling local ities for juvenile Atlantic croaker. /; is the number of individuals assayed from each locality. Region and locality Site Collection date n Standard length (mm) mean (SD) range Atlantic Delaware Bay, Delaware (DE) 38°48'N 75°10'W Oct 94 23 17.1(1.8) 13-20 Cape Fear River, North Carolina (NO 33°58'N 78°orw May 93 24 16.0(1.9) 13-19 Indian River. Florida (FL) 28'03'N 80°35'W Mar 94 22 19.4(2.2) 16-23 Gulf of Mexico Terrebonne Bay, Louisiana (LAl 29=10'N 90 30'W Jul 96 24 89.4(4.5) 79-98 (RFLP ) were used 1 ) to document the magnitude and spatial distribution of mtDNA variation in M. undulatus from U.S. Atlantic and Gulf of Mexico lo- calities, 2) to evaluate the integrity of Cape Hatteras, North Carolina, as a genetic stock boundary, and 3) to estimate levels of gene flow among Atlantic lo- calities to provide an improved basis for future deci- sions regarding coastwide management of this fish- ery resource. Methods Juvenile Atlantic croaker were collected from three U.S. Atlantic estuaries (Delaware Bay, Cape Fear River, and Indian River Lagoon) and one Gulf of Mexico estuary (Louisiana) (Table 1). Sampling lo- calities were chosen to represent the northern, cen- tral, and southern portions of the species" U.S. At- lantic coast range and to include areas north and south of the hypothesized stock boundary at Cape Hatteras, North Carolina. Genomic DNA extracts from individual croaker were obtained by using the Puregene DNA isolation kit (Centra Systems, Inc., Minneapolis. MN) and were used in the polymerase chain reaction (Saiki et al., 1988) to amplify two mtDNA regions: the pro- tein-coding ATP synthetase subunit 6 gene (ATPase 6 ) and the noncoding "D-loop" segment of the control region (Lankford, 1997). ATPase 6 was amplified by using the primers ATPase 6-L and ATPase 6-H (Quattro-^l. which yielded a 705-base-pair (bp) prod- uct. Primers L-Thr (5-3': AGC TCA GCG CCA GAG ^ Quattro, J. M. 1996. Department of Biological Sciences, Univ. South Carolina. Personal comniun. CGCCGGTCTTGTAAA)and 12SAR-H (5-3: ATA GTG GGG TAT CTA ATC CCA GTT) were used to amplify a 1540-bp product containing the entire D- loop region, tRNA-Pro and tRNA-Phe genes, and portions of the 12S rRNA and tRNA-Thr genes (Lee et al., 1995 ). Amplifications were performed in 50 ;uL reaction volumes according to Kocher et al. (1989). PCR products were digested with twelve restriction endonucleases: Hinfi, HaelW, Mspl, Pvull, Sau96I, Alul. Msel, /?,soI, Taql, Dclel, Nlalll, and Hhal (ob- tained from New England BioLabs, Beverly, MA) (Lankford, 1997). Variant RFLP patterns were con- firmed by means of repeat digestions and electro- phoresis on 39f NuSieve agarose gels. Fragment sizes were estimated by using molecular weight markers: pUC18-//at'III digest from Sigma-Aldrich Corp., St. Louis, MO, pBR322-M.spI digest from New England BioLabs, Beverly MA, and pBR322-Ss^Aa digest from New England BioLabs: and the ANAGEL software program (Mrazek and Spanova, 1992). Distinctive restriction-fragment patterns were identified by letter codes and subsequently combined to produce composite mtDNA haplotypes for indi- vidual fish. Nucleon diversity (/; ) was calculated for each locality (Nei and Tajima, 1981) with standard errors estimated according to Nei ( 1987 ). Nucleotide sequence diversity and nucleotide sequence diver- gence were calculated with the Restriction Enzyme Analysis Package (REAP, version 4.0) (McElroy et al., 1992). Frequency distributions of composite mtDNA haplotypes were tested for geogi'aphic homo- geneity by using a Monte Carlo chi-square simulation approach ( Roff and Bentzen, 1989) and a total of 10,000 random data resamplings. The integrity of Cape Hatteras as a genetic stock boundaiy was tested by comparing mtDNA haplotype frequencies north ( Dela- 886 Fishery Bulletin 97(4), 1999 ware) and south (North CaroHna and Florida) of this locaHty. Phylogenetic analysis was employed by using the character-state approach to construct a maximally parsimonious network relating individual haplotypes based on the number of restriction-site differences (Avise, 1994). The resulting network was examined for geographic structuring of haplotypes within and among sampling regions. Eopulation structure in M. undulatus was also evaluated by using nested analysis of molecular vari- ance (AM OVA [Excoffier et al., 1992] ). AMOVA input consisted of an Euclidean distance matrix contain- ing genetic distance values for all possible pairs of the 15 observed mtDNA haplotypes. Samples were stratified by locality (DE, NC, FL, and LA) and nested within region (Atlantic or GulD. Total genetic varia- tion was partitioned into three components: within geographic localities, among geographic localities within the Atlantic region, and between regions. Sig- nificance of variance components was tested by us- ing 10,000 random permutations to generate null distributions for each variance component. Gene flow among localities was estimated with Wright's (1943) island model, modified for mtDNA: Fg-j. = y2N.,m +1. Calculations employed 0gj as an approximation oiF^^. Results RFLP analysis of the ATPase 6 and D-loop regions revealed a total of 68 restriction sites. The average individual was scored for 62 sites and 264 nucleotide positions representing approximately 12% of the ATPase 6 and D-loop regions combined, or 1.6% of the entire mitochondrial genome. A total of 15 com- posite haplotypes was detected in M. undulatus (Table 2). Of the 93 individuals surveyed, 67 (72%) shared the same composite genotype (haplotype 1). Genetic distance among haplotypes averaged 0.17% for the pooled sample, with the most divergent haplotypes (10 and 12) differing from the common haplotype by only 0.75% and 1.0%, respectively. Nucleon diversity (/;) averaged 0.47 ±0.01 (mean ±SE) for the pooled sample, and ranged from h = 0.25 ±0.12 in DE to /z = 0.62 ±0.12 in LA. Nucleotide se- quence diversity also varied geographically, ranging from K = 0.07% in DE to ;r = 0.32% in LA. Net se- quence divergence (p) among localities was low, in- dicating that most of the observed mtDNA variation occurred within localities. Average sequence diver- gence for the pooled sample was p = 0.004% , with comparisons among Atlantic localities yielding slightly lower divergence values than comparisons involving Gulf and Atlantic samples. Croaker col- lected north versus south of Cape Hatteras were ge- Table 2 Geographic distribution (see Table 1 for specific locations) of composite mtDNA haplotypes among Atlantic croaker samples from Atlantic and Gulf of Me.xico localities. Let- ters denote mtDNA fragment patterns ( lowercase = ATPase 6; uppercase = D-loop) produced by digestion of PCR prod- ucts with the following polymorphic restriction endonu- cleases: ATPase 6: PvuU.Alul. Ddel. D-loop: Hiiifl, Haelll, Mspl, Msel, Taqh Nlalll, Hhal. Composite mtDNA haplotype Haplotype frequency DE NC FL LA Total 1. aaaAAAAAAA 20 16 16 1.5 67 2. bbaAAAAAAA 1 1 0 2 4 3. aaaAAAACAA 1 1 2 0 4 4. aaaAAAABAA 1 3 0 0 4 5. aaaAAAAAAB 0 1 0 0 1 6. aaaDAAAAAA 0 1 0 0 1 7. aaaABAAAAA 0 1 0 1 2 8. aaaBAAAAAA 0 0 3 0 3 9. aaaAAABAAA 0 0 1 0 1 10. aaaAAAABBC 0 0 0 1 11. aaaAAAAABA 0 0 0 1 12. aaaBBAAAAC 0 0 0 1 13. aaaAABAAAA 0 0 0 1 14. aabAAAAABA 0 0 0 1 15. aaaAAAAAAC 0 0 0 1 Totals 23 24 22 24 93 netically indistinguishable (p=-0.002%) whereas nucleotide divergence between regions (pooled Atlan- tic vs. Gulf) was 0.005%. The common mtDNA genotype (haplotype 1) was numerically dominant at all localities (Table 2). The remaining 14 haplotypes occurred at low (<10%) fre- quencies. Monte Carlo tests for homogeneity revealed no significant heterogeneity in mtDNA haplotype frequency distributions within the SAB (NC vs. FL, /-=11.3, p=0.140) or among all Atlantic localities (p=0.148). When NC and FL samples were pooled and compared to the DE sample, there was no indi- cation of heterogeneity among samples collected north and south of Cape Hatteras (^"=4.6, p=0.906). To further test for population structure within the Atlantic region, an additional 20 specimens from each Atlantic locality were screened for D-loop variation with the polymorphic enzymes Taql and Hhal. Nei- ther Taql (x''=3.45, p=0.562) nor Hhal (x-=1.95. p=0.682) haplotype frequencies exhibited significant heterogeneity among Atlantic localities. Given the lack of heterogeneity within the Atlan- tic region, composite haplotype frequencies were pooled across Atlantic localities and compared with the Gulf of Mexico sample. Monte Carlo simulations Lankford et al.; Population structure In Micropogonias undulatus 887 Table 3 Hierarchical nested analysis of molecular variance on genetic distance between Atlantic croaker mtDNA haplotypes Variance component df 0 statistic Variance % total P Among regions (Atlantic vs. Gulf) 1 0„ = 0.046 0.021 4.6 <0.01 Among localities within Atlantic 2 4)sf. < 0.001 <0.001 <0.1 0.10 Within localities 89 ST. = 0.046 0.431 95.4 0.03 revealed significant heterogeneity between Atlantic and Gulf samples (x2=24.36,p=0.020) (Table 2). Parsimony analysis indicated that geographically segregated haplotypes were generally closely related ( 1-2 restriction site changes) to the common haplotype (Fig. 1). AMOVA revealed that the majority (95.4%) of mtDNA variation in Atlantic croaker occurred within samples (Table 3). A significant portion (4.6%) was attributable to regional differences (Atlantic versus Gulf, P<0.01) but variation within the Atlantic region was unstructured (p=0.10) and accounted for <0.1% of the total genetic variance. 0gj values indicated a lack of geographic structure and relatively high gene flow among Atlantic localities. Gene flow among Atlantic localities was estimated as 23.4 (DE and NO, 16.8 (NC and FL), and 12.0 (DE and FL) effective female migrants per generation. O Atlantic DGulf • Atlantic & Gulf = 0.001 Nei's distance Figure 1 Parsimony network relating 15 mtDNA haplotypes observed in Atlantic croaker. Hatch marks drawn through each branch indi- cate the inferred number of restriction site differences among haplotypes. Branch lengths denote Nei's genetic distance between haplotypes (note scale). Discussion MtDNA analysis provided no evidence that M. undulatus is subdivided by Cape Hatteras into dis- crete genetic stocks. Frequency- and distance-based analyses both suggested a single, panmictic popula- tion of croaker on the U.S. Atlantic coast. Low levels of mtDNA divergence among Atlantic localities, al- though not statistically significant, were more con- sistent with a pattern of semi-isolation by distance rather than marked subdivision by Cape Hatteras. For example, AMOVA revealed that the two Atlantic localities exhibiting the least genetic divergence were DE and NC (0gT=OO21;p=O.89). NC and FL samples were slightly more divergent but not significantly so (cpg^=0.029; p= 0.180). Lack of population structure was perhaps best indicated by the lack of differen- tiation between DE and FL samples (g^=0.040; p=0.084>, two locations representing the distribu- tional limits of this species' range on the U.S. Atlantic coast. Kornfield and Bogdanowicz (1987) inferred pat- terns of gene flow from distributions of unique mtDNA haplotypes and their presumed precursors, predicting that under restricted gene flow, unique haplotypes should occur in the same population as their precursors. This was the case for M. undulatus at the regional (i.e. Atlantic versus Gulf) level, where unique Gulf haplotypes 10, 12, and 14 each had likely precursors that were confined to the Gulf region. Within the Atlantic region, however, unique haplo- types and their likely precursors occurred at dispar- ate localities. The occurrence of rare haplotypes at more than one Atlantic locality also implies that geno- types arising at one locality spread rapidly to other localities within the Atlantic. Combined with low lev- els of nucleotide divergence, these observations sup- port the hypothesis of contemporary gene flow among Atlantic localities. Gene flow estimates based upon Wright's island model are consistent with panmixia: values ranged from N^,m =12.0-23.4, well above the theoretical lower limit of Nj7i - 1 sufficient to pre- clude genetic divergence of populations by random drift (Slatkin, 1985). 888 Fishery Bulletin 97(4), 1999 Previous studies have reported geographic varia- tion in hfe history traits and population dynamics of Atlantic croaker found north and south of Cape Hatteras (White and Chittenden, 1977; Ross, 1988), suggesting that two groups of croaker with contrast- ing life histories overlap in the Cape Hatteras re- gion, with a mainly offshore group north of Cape Hatteras which displays greater size-at-age, greater longevity, lower annual mortality, and delayed matu- ration in relation to croaker south of Cape Hatteras. A more recent study by Barbieri et al. ( 1994a ) in the Chesapeake Bay concluded that life history varia- tion in M. iindulatits was ephemeral and that the two-stock hypothesis should be re-evaluated. The present mtDNA analysis supports previous sugges- tions by Ross ( 1988) and White and Chittenden ( 1977 ) that life history variation, when present, has an ecophenotypic basis. A review of A/, undulatus life history characteris- tics lends support to the panmixia hypothesis for the Atlantic region and provides several mechanisms by which trans-Hatteras gene flow may occur. Mark- recapture studies indicate extensive movement by M. undulatus along the U.S. Atlantic coast and thus potential for gene flow by means of adult migration (Pearson, 1932; Haven, 1959; Bearden, 1964; DeVries^ ). Larval dispersal could augment gene flow. Atlantic croaker larvae are abundant at outer-continental shelf locations (Govoni and Pietrafesa, 1994) and may spend more than 60 days in shelf waters before reach- ing estuaries (Warlen, 1980). Offshore spawning com- bined with an extended pelagic larval stage in shelf waters could provide ample opportunity for larval dispersal among Atlantic localities. Interestingly, spring recruits are occasionally reported in the mid- Atlantic Bight (MAB) estuaries as late as May and June (Haven, 1957; Chao and Musick, 1977) despite reports that spawning ceases in this area by late December (Morse, 1980; Barbieri et al., 1994b). Spring recruits to MAB estuaries may therefore be transported to mid-Atlantic estuaries from spawn- ing sites south of Cape Hatteras. Thorrold et al. (1997) found no significant differences in otolith chemistry of juvenile croaker from the Neuse River, North Carolina, and the Elizabeth River, Virginia, suggesting that immigrants to estuaries north and south of Cape Hatteras may originate from the same spawning area. The reproductive strategy of M. undulatus may also facilitate gene flow. Atlantic ' DeVries, D.A. 1986. InshoreAtlantic croaker tagging study In J. L. Ross, D. A. DeVries, J. H. Hawkins III. and J. B. Sullivan, Assessment of North Carolina commercial finfisheries, p. 3.31- 394. Completion report for Project 2-386-R. North (Carolina Department of Natural Resources and Community Develop- ment. Div. Marine Fisheries. Morehead City. NC. croaker are multiple spawners with indeterminate fecundity (Barbieri et al., 1994b). Because spawning occurs during a southerly coastal migration during fall, females could commence spawning within the MAB and spawn multiple times during their south- ward migration, effectively distributing maternal half-siblings (mtDNA clones) over an extensive geo- graphic range. MtDNA heterogeneity between Atlantic and Gulf of Mexico localities suggests that these regions sup- port separate populations of M. undulatus. The ob- served genetic break is consistent with a contempo- rary range discontinuity in southern Florida, where M. undulatus seldom occur south of Indian River on the Atlantic coast and are rarely encountered south of Tampa Bay on the Gulf coast (Nelson et al., 1991; Pattillo et al., 1997). Average sequence divergence between Atlantic and Gulf haplotypes (p=0.005'7f) was relatively low, however, suggesting that the phy- logenetic break between Atlantic and Gulf stocks may be less pronounced in M. undulatus than in other marine fishes (Bowen and Avise, 1990). This study represents the first attempt to identify population genetic structure in M. undulatus using DNA-level markers. The technique employed in our study proved sufficiently powerful to detect regional (Atlantic vs. Gulf of Mexico) population structure, but did not reveal structured genetic stocks of croaker within the Atlantic. The integrity of Cape Hatteras as a genetic stock boundary could be tested further by using fine-scale markers such as microsatellites, which have revealed genetic differentiation among fish populations that exhibit little mtDNA diver- gence. Alternatively, higher-resolution mutation- screening analysis or direct sequencing of the mito- chondrial D-loop region might provide greater reso- lution of geographic structure than the RFLP tech- nique employed here (Ong et al., 1996; Stabile et al., 1996). Although mtDNA analysis did not indicate discrete genetic stocks of croaker within the Atlantic region, harvest stocks (see Gauldie, 1988) worthy of man- agement consideration may exist. Because low lev- els of gene flow may produce mtDNA homogeneity between otherwise self-recruiting stocks, mtDNA is incapable of distinguishing between low {17c) and moderate {507f) amounts of mixing (Carvalho and Hauser, 1995 ). Delineation of harvest stocks will thus require more precise estimates of mixing than those obtainable from mtDNA analysis. Mark-recapture studies designed to quantify the level of adult mi- gration across Cape Hatteras, combined with otolith- based microchemical analyses to examine larval dis- persal patterns, could provide valuable information on the level of mixing between the MAB and South Lankford et al.: Population structure in Mtcropogontas undulatus 889 Atlantic Bight (SAB) areas and clarify the extent to which M. undulatus in these regions constitute self- recruiting groups. Acknowledgments We thank Mark E. Chittenden Jr and Steve W. Ross for valuable contributions to this project, and three anonymous reviewers for constructive comments. We are also grateful to Richard Paperno for logistical support during collecting trips to Indian River La- goon, Walt Pollard and Carolina Power and Light Company, Brunswick Generating Plant, for assis- tance during collecting trips to the Cape Fear River estuary, Michael Burger for providing specimens from the Gulf of Mexico, Joseph M. Quattro for donating oligonucleotide primers used to amplify the ATPase 6 fragment, and Ami E. Wilbur for instruction in the molecular laboratory. This research was supported by funding from the Office of Sea Grant, National Oceanic and Atmospheric Administration grant no. NA16RG0162-0L project no. R/F-5 to T. E. Targett, and by a University of Delaware, Marine Biology/Bio- chemistry Program Fellowship to T. E. Lankford Jr Literature cited Avise, J. C. 1994. Molecular markers, natural history and evolution. Chapman and Hall, New York, Slip. Barbieri, L. R., M. E. Chittenden Jr., and C. M. Jones. 1994a. Age, growth, and mortality of Atlantic croaker, Micropogonias undulatus. in the Chesapeake Bay region, with a discussion of apparent geographic changes in popu- lation dynamics. Fish. Bull. 92:1-12. Barbieri, L. R., M. E. Chittenden Jr., and S. K. Lowerre-Barbieri. 1994b. Maturity, spawning, and ovarian cycle of Atlantic croaker, Micropogonias undulatus. in the Chesapeake Bay and adjacent coastal waters. Fish. Bull. 92:671-68.5. Bearden, C. M. 1964. Distribution and abundance of Atlantic croaker. Micropogonias undulatus in South Carolina. Contrib. Bears Bluff Lab, 40, 2.3 p. Bowen, B. W., and J. C. Avise. 1990. Genetic structure of Atlantic and Gulf of Mexico popu- lations of sea bass, menhaden, and sturgeon: influence of zoogeographic factors and life-history patterns. Mar Biol. 107:371-381. Carvalho, G. R., and L. Hauser. 1995. Molecular genetics and the stock concept in fisheries. In G. R. Carvalho and T. J. Pitcher (eds.l, Mo- lecular genetics in fisheries, p. 55-79. Chapman and Hall, London, Chao, L. N., and J. A. Musick. 1977. Life history, feeding habits, and functional morphol- ogy of juvenile sciaenid fishes in the York River estuary. VA Fish. Bull, 75:67.5-702. Excoffier, L., P. E. Smouse, and J. M. Quattro. 1992. Analysis of molecular variance inferred from metric distances among DNA haplotypes: application to human mito- chondrial DNA restriction data. Genetics 131:479^91. Gauldie, R. W. 1988. Tagging and genetically isolated stocks offish: a test of one stock hypothesis and the development of another. J. Appl. Ichthyol, 4:168-173. Govoni, J. J., and L. J. Pietrafesa. 1994. Eulerian views of layered water currents, vertical distribution of some larval fishes, and inferred advective transport over the continental shelf off North Carolina, USA, in winter Fish. Oceanogr. 3:120-132. Haven, D. S. 1957. Distribution, growth, and availability of juvenile croaker. Micropogon undulatus. in Virginia. Ecology 38:88-97, 1959. Migration of the croaker, Micropogon undulatus. Copeia 1959:2.5-30. Kline, L. L., and H. Speir. 1993. Proceedings of a workshop on spot (Leiostomus xan- thurus) and Atlantic croaker i Micropogonias undulatus). Special Rep, 25 of the Atlantic States Marine Fisheries Commission. Washington, D.C.. 160 p. Kocher, T. D., W. K. Thomas, A. Meyer, S. V. Edwards, S. Paabo, F. X. Villablanca, and A. C. Wilson. 1989. Dynamics of mitochondrial DNA evolution in animals: amplification and sequencing with conserved primers. Proc. Nat. Acad. Sci. USA 86:6,196-6,200. Kornfield, I., and S. M. Bogdanowicz. 1987. Differentiation of mitochondrial DNA in Atlantic her- ring. Clupea harengus. Fish, Bull, 85:561-568. Lankford, T. E., Jr. 1997. Estuarine recruitment processes and stock structure in Atlantic croaker, Micropogonias undulatus (Linnaeus). Ph.D. diss.. LIniv. Delaware, Lewes. DE, 156 p. Lee, W. J., J. Conroy, W. Hunting Howell, and T. D. Kocher. 1995. Structure and evolution of teleost control regions. J. Mol. Evol. 41:54-66. McElroy, D., P. Moran, E. Bermingham, and I. Kornfield. 1992. REAP: the restriction enzyme analysis package. J. Hered. 83:157-158. Morse, W. W. 1980. Maturity, spawning and fecundity of Atlantic croaker, Micropogonias undulatus occurring north of Cape Hatteras, North Carolina. Fish. Bull. 78:190-195. Mrazek, J., and A. Spanova. 1992. ANAGEL. Comput, Appl, Biosci. 8:524. Nei. M. 1987. Molecular evolutionary genetics. Columbia Univ. Press. New ^xirk, NY, 512 p. Nei, M., and F. Tajima. 1981. DNA polymorphism detectable by restriction endonu- cleases. Genetics 97:145-163. Nelson, D. M., E. A. Irlandi, L. R. Settle, M. E. Monaco, and L. C. Coston-Clements. 1991. Distribution and abundance of fishes and inverte- brates in southeast estuaries. ELMR Rep. 9. NOAA/NOS Strategic Env. Assessments Div.. Rockville. MD. 177 p. Ong, T., J. Stabile, L Wirgin, and J. R. Waldman. 1996. Genetic divergence between Acipcnscr o.xyrinchus o.xyrinchus and A. o. desotoi as assessed by mitochondrial DNA sequencing analysis. Copeia 1996:464-469, Pattillo. M. E., T. E. Czapla, D. M. Nelson, and M. E. Monaco. 1997. Distribution and abundance of fishes and inverte- brates ill Gulf of Me.xico estuaries. II: species life history 890 Fishery Bulletin 97(4), 1999 summaries. ELMR Rep. 11. NOAA/NOS Strategic Env. Assessments Div.. Silver Spring. MD, 377 p. Pearson, J. C. 1932. Winter trawl fishery off the Virginia and North Caro- lina coasts. U.S. Bur Fish. Invest. Rep. 10, 31 p. Roff, D. A., and P. Bentzen. 1989. The statistical analysis of mitochondrial DNA poly- morphisms: X" and the problem of small samples. Mol. Biol. Evol. 6:539-545. Ross, S. W. -1988. Age, growth, and mortality of Atlantic croaker in North Carolina, with comments on population dynamics. Trans. Am. Fish. See. 117:461-473. Saiki, R. K., D. H. Gelfand, S. Stoffel, S. J. Scharf, R. Higuchi, G. T. Horn, K. B. MuUis. and H. A. Erlich. 1988. Primer-directed enzymatic amplification of DNA with a thermostable DNA polymerase. Science 239 (Wash. D.C.):487-491. Slatkin, M. 1985. Gene flow and the geographic structure of natural populations. Science 236 (Wash. D.C.):787-792. Stabile, J., J. R. Waldman, F. Parauka, and I. Wirgin. 1996. Stock structure and homing fidelity in gulf of Mexico sturgeon {Acipenser oxyrinchus desotoi) based on restric- tion fragment length polymorphism and sequence analy- sis of mitochondrial DNA. Genetics 144:767-775. Stone, S. L., T. A. Lowery, J. D. Field, C. D. Williams, D. M. Nelson, S. H. Jury, M. E. Monaco, and L. Andreasen. 1994. Distribution and abundance of fishes and inverte- brates in Mid-Atlantic estuaries. ELMR Rep. 12. NOAA/ NOS Strategic Env. Assessments Div, Silver Spring, MD. 280 p. Thorrold, S. R., C. M. Jones, and S. E. Campana. 1997. Response of otolith microchemistry to environmen- tal variations experienced by larval and juvenile Atlantic croaker (Micropogonias undulatiis). Limnol. Oceanogr. 42:102-111. Warlen, S. M. 1980. Age and growth of lar\'ae and spawning time of At- lantic croaker in North Carolina. Proc. Annu. Conf South- east Assoc. Fish Wildl. Agencies 34:204-214. White, M. L., and M. E. Chittenden Jr. 1977. Age determination, reproduction, and population dynamics ofthe Atlantic croaker, Micropogon/as undulatus. Fish. Bull. 75:109-123. Wilk, S. J., and H. E. Austin. 1981. Proceedings of the sciaenid assessment workshop, 17 September 1981, Stony Brook, NY. Virginia Inst. Marine Science. Gloucester Point. VA. 29 p. Wright, S. 1943. Isolation by distance. Genetics 28:114-138. 891 Abstract.— The age and growth of dol- phin, Coryphaena hippurus. has been determined in wild specimens between 14.4 and 124 era fork length (FL) caught off Majorca (Balearic Islands, western Mediterranean ). Also, several methods have been applied to validate the re- sults obtained and to assess the aging techniques presently used. In accor- dance with the sexual dimorphism of this species, the length-weight rela- tionship showed negative allometry in females and isometry in males. The high correlation between increment counts and known age in 75 reared lar- vae from 0 to 38 days old (number of increments = 0. 3894-1-0. 9846-days of age; r=0.98> indicated that the daily deposition of growth increments in sag- ittal otoliths and regular incremental formation begins on day one. This cor- relation validated the use of otoliths in the aging of 176 juvenile specimens between 16.5 and 58.5 cm FL, in which the minimum and maximum ages ob- served were 47 and 176 days, respec- tively. Nevertheless, the progressive increase in complexity with the onto- genic development of these structures suggests that this method may under- estimate age in adult fish, probably owing to the loss of marginal zones of the otolith during the polishing process. Thus, from 150 specimens between 65 and 124 cm FL, the interpretation of annuli from scales gave ages up to 3 years old, whereas readings of otoliths in 36 specimens of the same size inter- val gave ages between 0 and 1 year old. Growth parameters were calculated from the age-length relationships of otolith and scale readings in juvenile and adult fish, respectively, and from the analysis of monthly length-fre- quency distributions, obtained in 1990 and 1991 during the exploitation of ju- veniles from 18 to 70 cm FL (n =4084 1. The values obtained for L„ ranged from 72.4 cm FL in unsexed juveniles and 1 10.0 cm FL in females both juvenile and adult, whereas k ranged from 1.6/year in juvenile and adult females and 2.5/year in unsexed juveniles. Although the re- sults obtained were quite different ow- ing to the different length and age range considered, similar results were obtained by comparing the growth per- formance index (F), and showed the rapid growth of the species. Otolith microstructure, age, and growth patterns of dolphin, Coryphaena hippurus, in the western Mediterranean Enric Massuti I.E O , Centre Oceanografic de les Balears Moll de Ponent s/n, Apdo 291 07080 Palma de Mallorca Spam E-mail address emassuti g'dgpesca caibes Beatriz Morales-Nin Joan Moranta C-SI C./U I B ■ Institut Mediterrani d'Estudis Avancats Campus Universitan 07071 Palma de Mallorca, Spam Manuscript accepted 14 April 1999. Fish Bull. 97:891-899 (1999). Dolphin (Coryphaena hippurus Lin- naeus, 1758) (Pisces: Coryphaenidae) is an epipelagic, top-level predator species widely distributed in the tropical regions of every ocean (Palko et al., 1982). It migrates widely, and in warmer months in- creases its range to subtropical ar- eas. Thus, it occurs seasonally in the Mediterranean from spring to au- tumn, when surface waters reach temperatures above 16-18°C (Mas- suti and Morales-Nin, 1995); mature specimens are occasionally caught in the swordfish (Xiphias gladius) fishery with surface longlines at this time, and immature juvenile fish are exploited by small-scale fisheries that deploy surrounding purse nets around fish aggregation devices. Dolphin age and growth have been determined by means of scale annuli (Beardsley, 1967; Rose and Hassler, 1968), daily growth incre- ments in otoliths (Oxenford and Hunte, 1983; Uchiyama et al., 1986), and modal progression analy- sis (Wang, 1979; Murray, 1985). In addition, some growth studies have been made of fish of known age reared in captivity (e.g. Hassler and Hogarth, 1977; Ostrowski et al., 1989; Benetti et al., 1995b). In the Mediterranean, available growth data on this species are related to morphometric relationships (Lozano- Cabo, 1961; Bannister, 1976) and rearing experiments ( Rehones et al., 1992;Ayarietal., 1995). The objective of this paper was to determine, for the first time in the Mediterranean, the age and growth patterns (including juvenile and adult phases) of C. hippurus. To support age and gi'owth parameters estimated from readings of scales and otoliths, direct validation by means of rearing experiments and indirect validation based on length- frequency analysis were also ap- plied. Otolith microstructure was analyzed to determine increment patterns and morphological changes during growth and to obtain more precise increment counts. Material and methods From May to November 1990 and 1991, fork lengths (FL) of adult specimens (/; = 150) were measured from longline catches. Sex was de- termined on the basis of the sexual dimorphism of the species (Palko et 892 Fishery Bulletin 97(4), 1999 al., 1982). Scales were collected above the lateral line, at the level of the pectoral fin, cleaned with 5% KOH, and mounted on slides. Interpretation of scale rings was carried out with a projector at lOx magnifica- tion by using the method described by Beardsley (1967). The scarcity of adult fish prevented the use of age validation methods; therefore the annual na- ture of the rings was established from validated ages from other areas (Beardsley, 1967; Rose and Hassler, 1968). Juvenile fish (n=4084) were measured from 65 samplings made between August and December of the same years from small-scale fishery catches. Total weight and sex were recorded for subsamples (adults n =68, juveniles n=282) selected to cover the size distribution of both sexes. Sagittal otoliths were removed, cleaned, and stored in distilled water. Un- broken otoliths («=212) were embedded in epoxy resin and polished with a graded series of sandpa- per and, finally, with 0.3-|im alumina paste. To im- prove clarity, otoliths were moistened with immer- sion oil for several hours before reading. Otoliths have a complex structure, with convex sides, pronounced rostrum and antirostrum, and a demarcated sulcus that terminates in a V-shaped excisural notch (Oxenford and Hunte, 1983). We se- lected the dorsal side of the otolith to obtain a sec- tion with a complete increment sequence. Otoliths were read under a light microscope coupled to a high- resolution video camera and monitor system. Growth increments were counted from the core to the edge of the pararostrum (Oxenford and Hunte, 1983). In- cremental counts were made by beginning at the first clearly defined mark that encircled the primordium, which defines the outer edge of the nucleus. Growth increments in juvenile fish otoliths were enumerated along a single axis. In adult fish it was necessary, how- ever, to follow a circuitous path to complete a set of counts, following prominent increments laterally until an area with clear increments was found. Each otolith was read independently by two of the authors; results were accepted only if the readings were coincident or their difference was less than 57i . Growth increments were counted at 500x and verified at lOOOx. Microstruc- tural growth increments were studied on fish of 22^3 cm FL («=28) with a scanning electron microscope (SEM). To highlight increment zones, the polished sec- tions were etched for 30 seconds with a 0.1 N HCl solu- tion, or for 60-90 seconds with a 0.2 M EDTA solution. Von Bertalanffy growth parameters were esti- mated from age-length relationships obtained from scale and otolith readings by using the FISHPARM program (Prageretal., 1987). These parameters were validated by an independent length-based method by using the ELEFAN I procedure included in the ELEFAN software package (Gayanillo et al.M. Be- cause L^ and k are inversely correlated, the growth performance index (0 = 21ogL^ log^) was employed to compare growth rates (Munro and Pauly, 1983). The temporal relation between ring number and age was determined by using larvae of known age, obtained from eggs spawned in captivity. Larvae hatched on 25 July 1994 and were reared through the juvenile stage up to 38 days, according to the method described by Kraul (1989). Water tempera- tures during the experiment ranged from 26° to 27°C, within the range of water surface temperature found around the Balearic Islands from June to August, when the species spawns (at 24-27°C; Massuti and Morales-Nin, 1995). The presence of early larval stages has also been reported in the study area dur- ing this period (Alemany and Massuti, 1998). To establish the timing of first ring deposition and early otolith growth, larvae were sampled daily dur- ing the first 15 days. Afterwards, three samples were taken at intervals of 7-8 days until the juveniles were 38 days old. Larvae were preserved in buffered alco- hol. The standard length (SL) of each larva was mea- sured, and sagittal otoliths (n=93) were removed, cleaned, and mounted on slides with epoxy resin. The otoliths were read with an image analysis system connected to a microscope. An iterative semiauto- matic system was employed for detection of incre- ments, measurement of otolith radius, and age at- tribution. Increment clarity was improved by using polarized light, and enhanced digitized images were used for interpretation. Each otolith was read at least twice at lOOOx magnification with an oil lens. Daily periodicity of increments in reared specimens was validated by linear regression analysis of the relation between counts of otolith increments and days. The intercept and the slope were examined with Students <-test. This method was also used to test the isometry in the allometric indices obtained by applying exponential regression equations to the length-weight relationships. Results The parameters of the length-weight relationship for females, males, and the whole population, including unsexed specimens, are shown in Table 1. Isometric growth was found in males (Ntest; P>0.05l, but for females and the population as a whole, a negative allometry was found (^test; P<0.01). Gayanillo. F.C., M.Soriano, and D.Paulv- 1988. A draft guide to the complete ELEFAN. ICLARM Software Project 2, 70 p. International Center for Living Aquatic Resources Management, Manila, Philippines. Massuti et al.: Otolith microstructure, age, and growth of Coryphaena hippurus 893 Table 1 Length-weight relationship parameters estimated by ex- ponential regression equations (,v = ax^\ between fork length (cm) and total weight (g) in Coiyphaena hippurus. n - number of specimens; SE = standard error of slope; r = correlation coefficient. Population n SE Females Males Total 192 154 350 0.0139 0.0092 0.0113 2.8983 3.0187 2.9605 0.0011 0.997 0.0018 0.996 0.0007 0.996 The sagittae of C hippurus are small in rela- tion to fish size and generally butterfly-shaped, although they display different structural patterns with ontogenic development. The larval otoliths are round and the development of the rostrum starts at 8 days (Fig. 1, A and B). The sulcal struc- ture is formed in 12-day-old larvae (Fig. lO; the rostrum and antirostrum were, on average, sepa- rated in 17-day-old larvae (Fig. ID). Otoliths prepared for SEM observation showed differences depending on the etching agent. The otoliths etched with EDTA had less clear growth increments than did otoliths etched with HCl. These differences might be due to high protein content in the otoliths, which collapsed when the aragonite crystals were completely removed with EDTA. Whole otoliths showed the presence of concentric laminations or ridges (Fig. 2A) which revealed that the core region grew by the deposi- tion of successive layers on the inner side. Sec- tions of otoliths showed the typical pattern of in- cremental and discontinuous units, which formed growth increments with variable widths depend- ing on the otolith area (Fig. 2, B and C). Some rhythmic patterns were also evident with 7-14 increment groupings (Fig. 2, B and D). Under the light microscope, otoliths revealed a pattern of alternating light and dark concentric rings surrounding a core, which was deposited at the earliest stages of development (Fig. 3A). The width of the increments varied from the core to the edge, with narrow increments near the center of the otolith and at the edge, whereas wider ones were laid down from the first to the third month of life (Fig. 3, B and C). These differences in increment width may reflect periods of differential gi'owth. Larger increment patterns, or bands composed of groups of 7-14—28 increments, were also observed (Fig. 3B), suggesting lunar growth rhythms. From 93 reared lar\'ae of known age, a total of 75 otoliths (81%) could be read under the light micro- Figure 1 Photomicrographs of sagittal otoliths from Coryphaena hippurus larvae reared in captivity: lA) 6-day-old larvae of 6.1 mm SL (scale bar=25 ^m); (Bl 8-day-old larvae of 8.5 mm SL (scale bar=29 /jm); (C) 12-day-old larvae of 8.8 mm SL (scale bar=34 ;/mi: (Dl 17-day-old larvae of 30 mm SL (scale bar=56 /jml. (E) Poorly defined increments laid down around the core (scale bar=90 ^m). scope. These larvae measured 3.8-65.0 mm SL, and their otolith radii ranged from 10.7 mm (at day 0, 894 Fishery Bulletin 97(4), 1999 *^^V^ kv>? #:# X t.^"""^-'*^^-i5 »*-^*u jjr.: Figure 2 Scanning electron micrographs o? Coryphacna Jiippurus sagittal otoliths. (A) Ridges on the surface of the central area of the otolith (scale bar=155 jjm): (Bl thin increments laid down in the central region of the otolith, showing rhythmic growth patterns of 7- and 14-increment groupings (scale bar=35 urn); iC) wider increments laid down in the outer region of the otolith (scale bar=10 urn i; iDi rhythmic growth pattern of 14 daily growth increments (scale bar=15 /imi. 71=1) to 20.6-23.9 mm (at day 36, «=3). In general, poorly defined concentric increments, consisting of an adjacent dark and a light zone, were laid down around the core (Fig. IE). In relation between age in days and number of increments, neither the .r-axis intercept nor the slope were significantly different from 0 and 1, respectively (^test;P>0.05):y = 0.3894 + 0.9846.r (;!=75, /•=0.98). From the juvenile fish, 125 otoliths were read (71%), 73 females between 19.5 and 58.5 cm FL and 52 males between 16.5 and 58 cm FL. Observed mini- mum and maximum ages were 47 and 176 daily growth increments. Of 36 adult fish otoliths studied with the light mici-oscope, only 15 (42% ) were able to be interpreted. A high percentage were rejected ow- ing to error in reading precision greater than 5% or to erosion of the marginal zone during the polishing process. In specimens between 67 and 117 cm FL, the minimum and maximum ages read were 189 and 362 daily growth increments, respectively. Scales from 139 adult specimens, measuring 65- 124 cm FL, could be interpreted (93% ). The results showed the presence of three age groups within the population: 67 1-yr-old specimens ranging from 65 to 110 cm FL( -t =87.95, SD=10.15), 61 2-yr-oIds mea- suring 73-120 cm FL ( .v =97.54, SD=10.95) and 11 3-yr-olds measuring 92-124 cm FL (.v=108.73, SD=10.17). The von Bertalanffy growth function was calcu- lated on the basis of the age-length relationships from daily growth rings in juvenile otoliths and annulaein adult scales (Fig. 4). Adult daily ages were not taken into account in these calculations because of observed underestimations in aging adults from Massutf et al : Otolith microstructure, age, and growth of Coryphaena hippurus 895 otoliths. Growth parameters for the whole popula- tion and by sex are shown in Table 2. Length-frequency distributions for the 1990 and 1991 fishing seasons, based upon sampling of a fish- ery on juveniles, are given in Figure 5. Fish caught in this fishery from August to December measured 18-70 cm FL. The monthly length distribution indi- cated a rapid increase in the mode, mean, and mini- mum and maximum sizes for the catch. Von Bertalanffy growth parameters for this fraction of the population were estimated for the 1990 and 1991 fishing seasons (Table 2). Discussion The length-weight relationships obtained, showing negative allometric growth in females and isometry in males, are in accordance with the sexual dimor- phism of C. hippurus. In this species there is a char- acteristic bullhead in males which, in the western Mediterranean, begins its development in individu- als around 50 cm FL (Massuti and Morales-Nin, 1997). Similar results are obtained in other areas (Palko et al., 1982; Chatterji and Ansari, 1985), with a negative allometry that is stronger in females than in males. Our study confirmed the daily nature of otolith growth increments in the first month of life for C. hippurus as reared fish. The strong correlation of mean sagittal counts to known fish age validated the use of otolith growth increments in the aging of ju- veniles up to 40 days old. Because regular incremen- tal formation began on day one, no adjustment was required to estimate age from incremental counts of sagittae from wild fish. Results suggest that the age of adult specimens is underestimated by the otolith method. This is clear when age-length relationships from daily growth rings in otoliths and from annuli in scales are plot- ted. For the same size interval, the otolith reading gave ages between 0 and 1 year, whereas interpreta- tion of annuli in scales of adult fish gave ages to 3 years. This might be due to methodological problems in otolith preparation and interpretation. Consequently, the use of scales seems to be the best aging method for adult fish fi-om the Mediterranean, ft-om the point of view both of preparation and of results obtained. The interpretation of juvenile fish age from daily growth increments seemed accurate, with a low per- centage of disagreement between readers. To test the precision of this method, juvenile birthdate distri- bution was calculated by subtracting age in days from the date of capture. This showed a long period with peaks in the second fortnight of June 1990 and the first fortnight of July 1991 (Fig. 6). This small varia- Figure 3 Light micrographs of Coryphaena hippurus sagittal otoliths. (A) Daily growth increments in the core region of the otolith (scale bar=15 ^mi; iBl narrow increments in the central region of the otolith, showing rhythmic growth "patterns of 7-, 14-, and 28- increment groupings (scale bar=25;/m): (C) wider increments in the area of the otolith edge (scale bar=30 ^m). tion between years might be due to changes in the spawning peak or to differential mortality (Campana 896 Fishery Bulletin 9 7(4), 1 999 Table 2 Von Bertalanffy growth parameters for Coryphaena hippurus calculated by different methods. The standard errors of each pa- rameter are in parentheses, n = number of specimens; L, = asymptotic length (cm FL); k = growth coefficient in l/yr; Rn = goodness-of-fit index; r = correlation coefficient; 0 = growth performance index. LFA is length-frequency analysis; S and O are scale and otolith readings. Population Sample Method Rn Juveniles Juveniles Females Males Whole 1990 1991 1990-91 1990-91 1990-91 LFA LFA SandO SandO SandO 2635 72.4 2.5 — 0.447 1449 74.8 2.5 — 0.304 132 110 1.56 0.008 — (2.62) (0.14) (0.02) 132 98.7 2.06 0.024 — (1.59) (0.19) (0.02) 264 102.4 1.90 0.023 — (1.56) (0.15) (0,02) 0.92 0.91 0.87 4.118 4.146 4.277 4.302 4.300 140 120 - "E 100 - o ^ 80 -H 60 - I 40 - 20 - 0^ and Jones, 1992). According to our results and those of Uchi- yama et al. ( 1986 ), the deposition of increments in otoliths begins on the first day of life. In addi- tion, ripe eggs hatch within 50- 60 hours after fertilization, de- pending on temperature (Palko et al., 1982). Thus, hatching dis- tribution can be compared with the spawning period of the spe- cies in the area, which is known from adult fish maturity data (Massuti and Morales-Nin, 1995, 1997). The agreement between these hatching dates and the above-mentioned data supports the daily ages determined from otoliths of wild juvenile fish. Another independent verifica- tion method uses link ft-equency analysis. The modal progression in juvenile catches showed a rapid gi'owth rate, which is reflected by the almost double increase in monthly mean length during the fishing season fi-om late August to early December This rapid growth is in accordance with results obtained in the Mediterranean and other areas with reared specimens (e.g. Hassler and Hogarth, 1977; Renones et al, 1992; Ayari et al., 1995; Benetti et al., 1995b), and with the estimates of growth rates between 3 and 10 cm per month, obtained from modal progression analysis on wild specimens in different areas (Wang, 1979; Oxenford and Hunte, 1983; Murray 1985). Growth parameters obtained by the two different methods used for the two population groups were quite different owing to the different length ranges + 0.5 1 Juvenile otoliths - Whole — I 1.5 2 Age (years) Adult scales ■ Females 2.5 3 ■ Adult otoliths Males 3.5 Figure 4 Age-length relationships and von Bertalanffy growth curves obtained from scale and otolith interpretation in Coryphaena hippurus adults and juveniles. considered. Nevertheless, similar results are ob- tained with the growth performance index. These val- ues are similar to those calculated from growth pa- rameters reported from scales of annuli by Beardsley ( 1967 ) and Rose and Hassler ( 1968), which were 4.27 and 4.19, respectively. Those calculated from growth parameters obtained by daily growth increments (Uchiyama et al., 1986) were slightly higher (4.52 for females and 4.63 for males), probably owing to the above-mentioned underestimation of age in adult fish with this method, the few adult specimens stud- ied, and the effect of temperature on growth and metabolism. Temperatures can also explain the slower growth rate of C hippurus in the western Massuti et al ; Otolith microstructure, age, and growth of Coryphaena hippurus 897 30 20 10 0 August 90 (n = 389) mean: 24.10; S.D.: 2.10; range: 21-34 LL 20 15 - 10 - 5 - 0 August 91 (n= 103) mean: 25,89; S.D,: 3.56; range: 18-34 LX 10 15 20 25 30 35 40 45 50 55 60 65 70 10 15 20 25 30 35 40 45 50 55 60 65 70 September 90 (n = 922) mean: 35.35; S.D.: 5.07; range: 23-52 15 September 91 (n=635) mean: 32.84; S.D.: 4.44; range: 23^5 10 15 20 25 30 35 40 45 50 55 60 65 70 10 15 20 25 30 35 40 45 50 55 60 65 70 2 '5 i 10 I) a. 5 0 October 90 (n = 404) mean: 40.74; S.D.: 4.56; range: 28-58 ig h.. 6] 4 2 0 October 91 (n = 600) mean: 42.28, S D.: 5.08; range: 30-61 10 15 20 25 30 35 40 45 50 55 60 65 70 10 15 20 25 30 35 40 45 50 55 60 65 70 Fork length (cm) Fork length (cm) November 90 (n = 904) ,5 , mean: 46.03; S.D: 4.23; range 37-70 10 5 0 20 15 10 5 0 November 91 (n= 111) mean: 47.32; S.D.: 5.53; range: 34-65 J. JJ... 10 15 20 25 30 35 40 45 50 55 60 65 70 December 90 (n = 16) 3Q mean. 48.94, S.D.: 3.82, range: 43-56 10 15 20 25 30 35 40 45 50 55 60 65 70 20 10 0 ^ yji 10 15 20 25 30 35 40 45 50 55 60 65 70 Fork length (cm) Figure 5 Monthly length-frequency distributions of juvenile Coryphaena hippurus caught off Majorca with the surrounding net during the 1990 and 1991 fishing seasons. Fork length in cm (fi=number of specimens; SD=standard deviationi. North Atlantic Ocean (Beardsley, 1967; Rose and Hassler, 1968), where a decreased feeding rate when water temperature falls below 23^C and a cessation of feeding at 18°C have been reported (Hassler and Hogarth, 1977). By contrast, in Hawaiian waters, where faster growth rates have been reported, this spe- cies feeds throughout the year and can be expected to grow continuously (Uchiyama et al., 1986). The growth-rate values obtained for C. hippurus are within the range reported from otolith microstructures by Brothers et al. (1983) for Atlantic bluefin tuna (Thunnus thynnus), another top-level predator char- acteristic of the western Mediterranean epipelagic eco- system. According to Benetti et al. ( 1995a ), C. hippurus, Scombridae species, and Isurus oxyrhinchus, another Mediterranean pelagic predator, have common factors 898 Fishery Bulletin 9 7(4), 1999 in environment and lifestyle which must strongly select for anatomi- cal, biochemical, and physiological adaptations necessary to achieve exceptionally high metabolic rates. Such selection leads to high so- matic and gonadal growth rates, rapid digestion, and quick repay- me|it of oxygen debts, all impor- tant functions in these active pe- lagic predators. Acknowledgments 50 1990 (/i=49) H 1* 1991 (/i=76) M5 May 16-31 May 1-15 Jun 16-30 Jun 1-15 Jul 16-31 Jul 1-15 Aug 16-31 Aug 1-15 Sep The authors wish to thank the staff of Centre Oceanografic de les Balears (I.E.O.) and the crews of FVs Es Form. Fiiria, Orion HI. and Virgen del Carmen II for the facilities provided, as well as col- leagues who participated in sampling. Special thanks are due to F. Doumenge (Institut Oceanographique Monaco) and O. Bourgeois, G. Dewavrin, and E. Poilpre (Pisciculture Marine Monaco S.A.M.) for col- laboration during the validation experiment. Also, R. L. Radtke (Hawaii Institute of Geophysics, Ha- waii), J. Uchiyama (Southwest Fisheries Center Honolulu Laboratory, Hawaii), and Z. Wang (Hawaii Institute of Geophysics, Hawaii) helped us in the initial steps of this study. J. M. Fortune and J. Biosca (Institut de Ciencies del Mar, Barcelona) are thanked for the SEM micrographs, and C. Rodgers for the revision of the English version of the manuscript. Literature cited Alemany, F., and E. Massuti. 1998. First record of larval stages ofCoryphaena hippurus (Pisces: Coryphaenidae) in the Mediterranean .Sea. Sci. Mar 62:181-184. Ayari, A., H. Ben Ouada, and B. Peyrou. 1995. Elevagede la doradecoryphene I Corvp/joena hippurus). Cah. Options Mediterr 16:12.'5-130. Bannister, J. V. 1976. The length-weight relationship, condition factor and gut contents of the dolphin-fish Coryphaena hippurus (L.) in the Mediterranean. J. Fish Biol. 9:335-338. Beardsley, G. L. 1967. Age, growth and reproduction of the dolphin. Coryphaena hippurus, in the Straits of Florida. Copeia 1967:441-451. Benetti, D. D., R. W. Brill, and S. A. Kraul Jr. 1995a. The standard metabolic rate of dolphin fish. J. Fish Biol. 46:987-996. Benetti, D. D., E. E. Iversen, and A. C. Ostrowski. 1995b. Growth rates of captive dolphin, Coryphaena hip- purus, in Hawaii. Fi.sh, Bull. 93:152-157. Birthdaie period Figure 6 Frequency distribution of back-calculated birthdates for Coryphaena hippurus ju- venile specimens in 1990 and 1991. Brothers, E. B., E. D. Prince, and D. W. Lee. 1983. Age and growth of young-of-the-year bluefin tuna, Thunnus thynnus, from otolith microstructure. In E, D. Prince and L. M. Pulos (eds.) Proceedings of the interna- tional workshop on age determination of oceanic pelagic fishes: tunas, billfishes, and sharks, p. 49-59. U.S. Dep. Commer. NOAA Tech. Rep. NMFS 8. Campana, S. E., and C. M. Jones. 1992. Analysis of otolith microstructure data. In D. K. Stevenson and S. E. Campana (eds.). Otolith microstruc- ture. Examination and analysis, p. 73-100. Can. Spec. Publ. Fish. Aquat- Sci. 117. Chatterji, A., and Z. A. Ansari. 1985. A note on the length-weight relationship in dolphin fish. Coryphaena hippurus L. Mahasagar 18:425-427. Hassler, W. W., and W. T. Hogarth. 1977. The growth and culture of dolphin, Coryphaena hippurus. in North Carolina. Aquaculture 12:115-122. Kraul, S. 1989. Review and current status of the aquaculture poten- tial for the mahimahi, Coryphaena hippurus. Aquacop. IFREMER. Actes de CoUoque 9:44.5-4.59. Lozano-Cabo, F. 1961. Biometria, biologiay pescade la llampuga( Corvp'iaena hippurus L.) de las Islas Baleares. Mem. R. Acad. Cienc. Exactas Fis. Nat. Madr Ser. Cienc. Nat. XXI. 93 p. Massuti, E., and B. Morales-Nin. 1995. Seasonality and reproduction of dolphin-fish (Cory- phaena hippurus) in the western Mediterranean. Sci. Mar. 59:357-364. 1997. Reproductive biology of dolphin-fish {Coryphaena hippurus L.) off the island of Majorca (western Mediter- ranean). Fish. Res. 30:57-65. Munro, J. L., and D. Pauly. 1983. A simple method for comparing the growth of fishes and invertebrates. Fishbyte 1:5-6. Murray, P. A. 1985. Growth and mortality in the dolphin-fish Coiyphaena hippurus caught off Saint Lucia, W.I. FAO Fish. Rep. 327:47-153. Massuti et al : Otolith microstructure, age, and growth of Coiyphaena hippurus 899 Ostrowski, A. C, C. Brownell, and E. O. Duerr. 1989. Growth and feeding rates of juvenile dolphins (Coryphaena hippurus) fed a practical diet through growout World Aquacult. 20:104-105. Oxenford, H. A., and W. Hunte. 1983. Age and growth of dolphin, Coryphaena hippurus, as determined by growth rings in otoliths. Fish. Bull. 84: 906-909. Palko, B. J., G. L. Beardsley, and W. J. Richards. 1982. Synopsis of the biological data on dolphin-fishes, Coryphaena hippurus Linnaeus and Coryphaena equiselis Linnaeus. FAO Fish. Synop. 130, 28 p. Prager, M. H., S. B. Saila, and C. W. Recksiek. 1987. FISHPARM: a microcomputer program for parameter estimation of nonlinear models in fishery science. Old Dominion Univ. Res. Found. Tech. Rep. 87, 37 p. Renones, O., E. Massuti, and B. Morales-Nin. 1992. Primeras experiencias de captura y mantenimiento en cautividad de llampuga (Coryphaena hippurus) en Mallorca. Inf Tec. Inst. Esp. Oceanogr. 123, 15 p. Rose, C. D., and W. W. Hassler. 1968. Age and growth of the dolphin, Coryphaena hippurus (Linnaeus), in North Carolina waters. Trans. Am. Fish. Soc. 97:271-276. Uchiyama, J. H., R. K. Burch, and S. A. Kraul Jr. 1986. Growth of dolphins. Coryphaena hippurus and C. equiselis, in Hawaiian waters as determined by daily in- crements on otoliths. Fish. Bull. 84:186-191. Wang, C. H. 1979. A study of population dynamics of dolphin-fish (Cory- phaena hippurus) in waters adjacent to Eastern Taiwan. Acta Oceanogr. Taiwan. 10:233-251. 900 Abstract.— The social structure and genealogical relationships of resident killer whale pods in Prince William Sound, Alaska, are inferred from asso- ciation analysis and direct observation. During 1984-95 a total of 2444 hours of observation were made and 36,009 photographs were taken of identifiable killer whales. Cole's association index and a point correlation coefficient in- dex were used to test the statistical sig- nificance and strength of associations between individuals, and a clustering procedure was used to delineate group structure. A total of 202 whales were grouped into 9 pods. Genealogical rela- tionships were inferred from the strength of bonds among pod members. Genealogical trees suggested that intrapod groups were matrilineal in structure. Splitting of one pod I AN podi was observed during the study; how- ever, there was no splitting of matri- lineal groups. Association patterns and inferred genealogies of resident l be. for be > ad and d > i propriate for an analysis of patterns of association between individuals because not all individuals were equally identifiable (see Bigg et al., 1990) as was es- pecially the case for cow-calf pairs, in which the of- ten well-marked mother was generally identified in more photographs than her usually indistinct young calves even though they always traveled together. Only CAI values for all years' combined were used to determine intrapod groups. CAI values were later calculated for the periods 1984-88 and 1989-95 and used to examine changes in bond strength between mothers and female offspring as the offspring ma- tured and produced their own calves. The relation- ship between age of males and the bond strength with their mothers was also examined with these CAI values. Groupings of individuals were identified from den- drograms constructed with an agglomerative aver- age single-link algorithm (Johnson, 1967; see Fig. 1 ) In this procedure, CAI values for all possible pairs of individuals were compared and the pair with the highest CAI was linked. Then the pair of unlinked individuals with the highest CAI was linked, or an unlinked individual with a higher mean CAI value with previously linked individuals was linked to that pair, and so on until the mean CAI dropped to 20''r. By this point, the vast majority of individuals had been linked into intrapod groups and we switched to an analysis with the point correlation coefficient (PCC): CAI SE = ad -be ~ i.b + d){e + d) (a + b){a + e) n{b + d){e + a) for be > ad and a > d. PCC ad - be The index was expressed as a percentage ranging from 100 to -100, with lOO'/r indicating that the joint number of occurrences of each whale equaled the number of occurrences of the least-photographed in- dividual. Zero percent indicated that the individu- als were randomly distributed among photographs, and -1009^ indicated that the two individuals were never photographed together. The statistical signifi- cance of CAI values was evaluated according to their standard errors using Student's t test (Cole, 1949). The CAI provides a measure of complete associa- tion ( rather than absolute association ) in that a value of 1009f occurs only when the'joint number of occur- rences equals the number of occurrences of the less frequently photographed individual (see Cole, 1949, for details and a review of association indices). An index of complete association was deemed more ap- [(a + b)(a + e)ib + d)(c + d)] where a represents the number of photographs con- taining one or more members of both groups, b and e represent the numbers containing members of only one or the other of the groups, and d represents the number containing no members of either group. The PCC index, a measure of absolute association, was used to examine associations among intrapod groups determined by CAI analysis because each group was considered equally identifiable. The PCC index varies from 1009f to -1009r, with 0 indicating a random distribution. An index of absolute associa- tion was deemed to be more appropriate for intrapod groups because all groups contained readily identi- fiable animals; therefore observation of one group in the absence of another group indicated that each group was traveling independently of one another. The relationship.s among intrapod groups were ex- amined from dendrograms constructed from PCC values with an agglomerative average single-link algorithm (see Fig. 2). Linkages with a positive PCC Matkin et al,; Association patterns of Orcinus orca 903 20 30 40 50 60 70 80 90 100 n CM m BD6 S37 ao B34 B27 BX B4g B41 B18 B07 808 B12 B23 B3S aog B01 B21 B16 B«S -68 71 0 76 . -91 15 89 2 . ■90 -66 -95 l«S B19 B17 B13 B<3 B3S B2D B14 -•^ *l -91 -66 -7J 54 • ^! -14 94 • -69 85 59 J) 77 ■76 -93 -76 -75 75 -18 • • 78 • • 56 -51 -57 u -74 -68 -72 -65 ■65 -95 -65 17 .77 -63 -87 53 -65 -81 -71 B1S B« B24 B22 B« B44 -69 . 12 I) • -82 -92 M il i2 37 . -70 -2 90 55 -65 38 • -68 -61 56 92 ABP99 ABOld AB219 ABieo AB19d I AB45C91) \0a 22 71 82 88 87 AB140 c AB24d AB15 AB40 ('88)d AB229 AB349 -16 -17 -72 -70 Jl -67 -77 -90 -79 84 -66 79 91 -87 *1 -54 98 -95 -95 -81 ■ -24 -88 -79 -80 -96 -58 -60 -84 -57 94 89 • • 93 40 -87 -75 -79 ■81 -71 -84 15 -79 55 98 -92 -37 -75 -91 -67 18 17 -26 -58 1 3 ■75 25 19 -5 41 4 26 -73 20 14 012 016 I 2 -69 1240 6 49 .10 -10 54 26 1 *5 78 AB28 AB269 AB279 I I AB44 (M) AB49 ('94) AB48 ('93) AB50 ('96) ABI79 ■61 ^61 0 W aoe B37 803 BW 327 AB1 AB1 (■88) Whale Designation Whale Deceased Known Date 0I Birth Known Relationship Proposed Relationship (Nun.. Br of valid photographic sequences in which animal appears Is at base of matnx column.) AB209 AB35d ABI39 AB43('88) AB06 9- 848 321 841 -66 B18 307 808 B13 8Z9 BS 809 801 B21 22 816 • B4S 49 B19 ' B17 1 813 4 BO 91 B35 84 B2D '3 B14 • 815 41 840 '6 824 59 822 ■ 84) 70 94, ABoad AB239 ABsecse) AB37 ('86) Aa)79 AB08 9 I AB41I ('86) AB12 AB18 Figure 1 Top: dendrogram illustrating intrapod groups and relationships among individuals with CAI values calculated at the population level. (Relationships among intrapod groups are shown in Fig. 2.) Middle: matrix of CAI values between pairs of individuals calculated at the pod level for AB pod. AB17 subpod. Numbers at the base of each column in the matrix are the number of valid photo sequences in which that individual appears. Bottom: inferred genealogical trees. 904 Fishery Bulletin 97(4), 1999 AB2S Subpod 60 B02 B25 B32 Boe B34 B07 AB17 B23 SubpodB09 B17 B14 B22 AB10B10 Al 103 001 ADS Pod D07 D11 005 012 ADie Die AK K02 Pod K06 N13 N34 N16 AN20 N17 Pod N20 N19 N23 N27 N29 N05 AN10 N06 Pod N09 N10 J02 J08 JOS J06 AJ J14 Pod J23 jie J18 J20 AE Pod E02 E11 E12 EOS E14 60 50 Point Correlation Association index (PCC) 40 30 20 AB25 10 -10 Subpod AB17 Subpod AB10 Subpod AlPod ADS Pod AD16 Pod 3 AKPod AN20 Pod AN10 Pod AJPod AEPod 50 40 30 20 10 -10 Figure 2 Dendrogram illustrating the relationships among intrapod groups (see top of Fig. 1 and Figs. 3-12) with values derived from PCC analysis. association value (PCC>0) were designated as pods (Fig. 2). Groups were designated as subpods rather than pods when PCC values between them did not J exceed zero but direct observation and photographic Matkin et aL: Association patterns of Orcinus orca 905 summaries of encounters indicated that designated subpods traveled together as a single pod over 50% of the time. Determination of sex and age Sexually mature males were differentiated from fe- males and immature males by their higher dorsal- fin height-to-width ratio (HWR), which typically ex- ceeds 1.4 by 15 years of age (Olesiuk et al., 1990). Mature females were identified when they gave birth and were accompanied by a new calf The sex of most juveniles could not be determined except in cases where the penis or the unique pigmentation pattern of the genital region was observed (Bigg et al., 1990; Olesiuk et al., 1990). Actual ages could be determined for whales born during the study on the basis of their birth year. The ages of whales that were immature at the beginning of the study were estimated when they were first seen on the basis of relative size of the whale and size of the dorsal fin (Bigg et al., 1990). The approximate year of birth for whales that matured during the study was estimated by subtracting the mean age of maturity ( 15 years for both sexes) from the year they matured. Females were considered to have matured in the year they gave birth to their first viable calf, and males in the year in which their dorsal fin at- tained an HWR of 1.4 (Bigg et al., 1990; Olesiuk et al., 1990). The year of birth of males that were sexu- ally but not physically mature at the start of the study was estimated by subtracting the mean age of physical maturity from the year their dorsal fin at- tained a HWR of 1.6-1.8, indicating physical matu- rity. Males that were physically mature and had a dorsal fin HWR of 1.6-1.8 at the beginning of the study were considered to be at least 21 years of age at that time. The minimum age of females that were mature at the beginning of the study was estimated by subtracting 15 years from the estimated birth year of their eldest offspring. This was a minimum esti- mate because a female's elder offspring may have died before the start of the study. Females that had not given birth for a decade or more were considered postreproductive (Olesiuk et al., 1990). Construction of genealogical trees Because statistical analysis indicated that whales associated in stable groups (or pods), CAI values were recalculated for all individuals within each pod and displayed in a matrix to show the relative strength -of association among pod members (Fig. 1). We pos- tulated that associations within pods reflected ge- nealogical relationships and used them in conjunc- tion with data on known relationships, sex and age data, and direct obsei-vations to infer genealogical relationships (Fig. 1). Possible maternal genealogical trees were con- structed in three steps (Bigg et al., 1990). Individu- als to be incorporated into the tree as offspring were selected beginning with those born during the study, followed by those that were juvenile at the start of the study, and finally by those that were mature at the start of the study. Second, their potential moth- ers were identified. All mature females in the same pod were considered, providing that they could have been at least 12 years (minimum age of maturity) older. An individual's own mature daughters were excluded as potential mothers. Third, the relative strength of bonds as indicated by CAI values between the individual and all its potential mothers was ex- amined. The potential mother with which the indi- vidual was most closely bonded was assumed to be its mother. An individual not strongly bonded to any potential mother was not assigned a mother. Matri- ces were cross-checked to insure that mother-off- spring assignments created sibling groups that dem- onstrated reasonable linkage by CAI values. Direct observation was used to supplement statistical analy- sis to construct genealogies in instances where indi- viduals and groups were less frequently photo- graphed and where there was some ambiguity in the numerical analysis. Results A total of 2444 hours of direct observation of whales was logged from 1984 to 1995 (Table 1), during which 36,009 frames of film were exposed that were suit- able for use in statistical analysis of association pat- terns (more than one individual appeared in the pho- tographic sequence). A total of 202 whales photo- graphed between 1984 and 1995 were included in the association analysis. According to direct field observations and inspection of photographs, these had previously been assigned to 9 pods (Table 2; Heise et al., 1992). Another 158 whales were tentatively grouped into 5 pods by field observation and visual inspection of photographs (pod AX, 54 whales; pod AY, 11 whales; pod AS, 17 whales; pod AF, 48 whales; pod AG, 28 whales). These pods were not included in the association analysis because there were insuffi- cient photographs. The individuals listed in Table 2 represent the cumulative membership of pods over all years of the study. In all pods and most groups, the numbers of individuals varied over the course of the study as members died or were born during the study. 906 Fishery Bulletin 97(4), 1999 Table 1 Hours of direct observation and number of usabl s film fr ames by year Year 1984 1985 1986 1987 1988 1989 1990 1991 1992 1993 1994 1995 Total Frames Hours 6,076 307 1.284 154 2,967 127 1.326 91 1,400 3,549 92 268 5,940 423 3,701 246 3,641 255 2,333 127 1,662 120 2,130 36,009 234 2444 Statistical analysis indicated that individual whales did not associate randomly with one another. For individuals seen together at least once, 38.5% of all CAI values from the initial analysis of all indi- viduals in all years were significantly less than 0 at a 5% confidence level (30.1% at a 1% level) and 27.5% were significantly greater than 0 at a 5% level (23.2% at a 1% level) (see Cole, 1949). Because these values are far greater than expected owing to chance, we rejected the null hypothesis that association patterns were random. The various animals positively associ- ated with a particular individual tended to be posi- tively associated themselves. These groups were also indicated by the agglomerative cluster analysis. A dendrogram showing associations among all 202 individuals included in the analyses was too large to exhibit here as a single figure; it appears split into pods along with matrices of CAI values and proposed genealogical trees for each pod (Figs. 1, and 3-12) In nearly all cases, intrapod groups of individuals linked by association analysis at greater than 20% CAI reflected groupings established by direct obser- vation. Direct observation also indicated that these intrapod groups virtually always traveled together. Most intrapod groups were centered around a repro- ductive or postreproductive female, which led us to suspect that these groups reflected matrilineal ge- nealogies. However, in four cases single males (AE14, J02, N19, and DOl) were not joined to other intrapod groups (CAI>20%) and in two cases pairs of males (AD02 and AD12, and AB02 and AB29) were linked to each other but not to other intrapod groups (CAI>20%). The nine pods examined contained 48 intrapod groups, including male singles and pairs ( Figs. 1 and 3-12). With the exception of AB pod, pods com- prised 1-9 intrapod groups. Two pods (AI and AD 16) contained only one intrapod group. The three subpods of AB pod contained a total of 12 intrapod groups. Intrapod groups were composed of 1-9 individuals. Patterns of association between intrapod groups, as indicated by statistical analysis, generally sup- ported the pods constructed by direct observation. Forty-six of the 48 intrapod groups (190 of 202 indi- viduals) were linked at the zero level of association into their respective pods (Fig. 2). There were two Table 2 Pods and individual whales used in this analysis. Pod Cumulative membership AB AB01-AB51 AI AI01-AI07 AJ AJ01-AJ38 AE AE01-AE20 AK AK01-AK14 ANIO AN01-AN03, AN05-AN12, AN35,AN38, AN40- AN41,AN45-AN51 AN20 AN04, AN13-AN34, AN36-AN37, AN39, AN42- AN44 ADOS AD01-AD12, AD19,AD21-AD27 AD16 AD13-18,AD20 exceptions; repeated direct observations were used to clarify separations into intrapod groups. First, the ABIO subpod was linked in the dendrogram with AI pod (PCC=4) before it was linked with the other subpods of AB pod. Second, AK pod and AD16 pod formed distinct clusters as expected but were joined at the PCC = 8 level, which was greater than the zero level adopted to define the other pods. On the basis of direct observation, ANIO pod and AN20 pod were considered a single pod (AN pod) until 1989, when they began traveling separately a ma- jority of the time. AN whales were encountered on 46 occasions during 1984-88; the ANIO and AN20 groups were observed together in 35 (76%) of these encounters. In contrast, during 1989-95 they were observed together in only 3 of 65 (5%) encounters where AN whales were present. After 1992 they were not seen traveling together and were thus designated as separate pods. The association analysis supported their designation as separate pods (Fig. 1). AD pod was also considered a single pod early in the study on the basis of direct observation during a few en- counters. From more recent direct observation and the results of the association analysis, it is now des- ignated as two pods (AD5 and AD 16 pods). Matkin et aL; Association patterns of Orcinus orca 907 20 i 30 40 50 60 70 80 90 100 BQ2B29B25B46B39B30B31B38B32B47B33B42 563 12 -23 427 -10 1161 •52 65 238 AB329 ■36 -44 31 17 742 ■34 -41 -12 6 372 AB30 AB339 AB42('88) -22 -28 -19 -29 -2 496 -45 27 -9 12 -10 14 42 591 AB47 ('92) -16 -56 -42 -76 -51 2 -2 -54 lllg AB02d ••AB259 AB3,19 10 -65 -75 -68 -72 76 244 AB29C5 -19 -65 ^1 -72 -47 8 9 -53 37 25 979 AB38 ('87) AB399('87) AB46 ('92) AB51 Cgfi) -10 B02 74 B29 35 B25 • B46 -43 B39 26 B30 46 B31 -39 B38 25 B32 • B47 22 B33 181 B42 AB1 Whale Designation AB1 Whale Deceased (•88) Known Date of Birth Known Relationship ProDosed RalationAhin (Number of valid prwtographtc sequences in which animsl appears Is at boss of mathx column.) Figure 3 Top: dendrogram illustrating intrapod groups and relationships among individuals with CAI values calculated at the population level. ( Relationships among intrapod groups shown in Fig. 2.) Middle: matrix showing CAI values between pairs of individuals calculated at the pod level for AB pod, AB25 subpod. Numbers at the base of each column in the matrix are the number of valid photo sequences in which that individual appears. Bottom: in- ferred genealogical trees. On the basis of lack of linkage in the PCC dendro- grams above the zero level, AB pod was divided into three subpods (ABIO, AB17, AB25), although prior to 1994, direct field observations indicated that these .subpods very rarely traveled separately. However, even when traveling together, the subpods typically traveled as cohesive units within the pod. Since 1994, direct observation has indicated that the AB25 sub- pod has split off and travels mainly with AJ pod. Other than those established for AB pod, no other subpods were identified in the study. To explain association patterns within intrapod groups, we postulated that these represented matri- lineal groups and that genealogical relationships 908 Fishery Bulletin 97(4), 1999 20 30 40 50 60 70 80 90 100 B10 B11 B04 BOS 826 ■12 703 -37 78S -45 -28 -23 589 BIO B11 B04 BOS ABlOg ABOSd AB04C^ AB11C5 ABl Whale Designation ABt Whale Deceased (■88) Known Date of Birth Known Relationship Proposed Relationship (Number of valic In which anima matrix column.) photographic se<|uencas appears Is at base of Figure 4 Top: dendrogram illustrating intrapod groups and relationships among individuals with CAI values calculated at the population level. (Relationships among intrapod groups shown in Fig. 12.) Middle: matrix showing CAI values between pairs of indi- viduals calculated at the pod level for AB pod, ABIO subpod. Numbers at the base of each column in the matrix are the number of valid photo sequences in which that individual appears. Bottom: inferred ge- nealogical trees. could be deduced by examination of bond strengths within these groups. CAI values were calculated be- tween all individuals of each pod and dendrograms of the CAI values were constructed to reveal the structure of associations between the members of each pod (Figs. 1, and 3-12). The individuals in the nine pods examined in the second CAI analysis were placed in 39 postulated Table 3 Cole's associa male offspring ion index (CAI) values for mothers and fe- that produced their first calf in 1987 or later Reproductive offspring Year of first calf Mother CAI 1984-1988 CAI 1989-1995 AJ3 1990 AJ8 28 13 AJ4 1994 AJ8 57 20 AJ13 1992 AJ14 40 33 AK7 1993 AK6 51 24 ANIO 1987 AN9 22 -20 ANll 1992 AN9 30 23 AN26 1990 AN23 53 34 AN31 1990 AN20 28 19 AN35 1988 AN9 29 2 maternal genealogical trees on the basis of linkages established in association analysis. An additional four newborn calves were observed in 1996 and placed in the genealogical trees on the basis of field observa- tions only. Genealogical trees were constructed by first establishing all known mother-offspring rela- tionships from field observations. There were 58 off- spring born during the study that appeared in the photographs used for association analysis. In all but one case (AB41 and mother AB8), the CAI values be- tween offspring and their known mothers was higher than for any other pairings of individuals. Direct ob- servation also indicated that offspring maintained their strongest bonds with their known mothers throughout the study period. The three offspring born at the be- ginning of the study were still most strongly bonded to their mothers after 12 years. These were mother-off- spring pairs AI3 and AI4 (CAI=27); AK6 and AK8 ( CAI=34 ); and AE 1 1 and AE 13 ( CAI=46 ), The strength of the bond between females and their mothers declined in the nine cases in which females became reproductive during the study and their mothers remained alive throughout the study (Table 3), A paired t test of CAI values for 1984-88 and CAI values for 1989-95 showed that the tendency for as- sociations to decline after female offspring become reproductive was statistically significant (^=4,88, P=0.0012). There were 31 juveniles (estimated age 10 years or less in 1984) at the beginning of the study. Asso- ciation analysis indicated that all but five of these whales remained most closely bonded to the whales judged by direct observation to be their mothers. Three of these exceptions were females that produced calves (AK7 and calf AK12: AN8 and calves AN41 Matkin et a\ Association patterns of Orcinus orca 909 20 30 40 50 60 70 80 90 100 037 173 CB2 60 69 51 36 59 DCS ■1 001 3 6 53 003 •59 .74 2 93 AD02(5 ADOSQ DID 49 40 50 121 008 2« 12 13 29 -73 -85 -71 177 D23 -70 -87 -4\ AD12(? ADOed AD21 CgO) AD25 CSS) AP049 100 27 004 -4 8 6 -72 33 44 57 6 100 164 Oil ■52 -39 -16 019 ■35 -18 -46 026 ■2 ■18 ■17 005 -76 -76 D2S -75 -81 ■90 006 ■74 021 002 ■77 ■86 ■82 ■38 42 ■17 ■28 ■20 200 ■30 55 193 ■20 ■28 15 35 47 92 ADOld AD03(5 AD10 ■78 ■67 -91 207 -90 -83 75 57 ADII9 AB1 AB1 (■88) Whale Oe«lgnat)on Whale Decsased Known Date o( Birth Known Relationship Proposed Relationship (Number of valid photographic sequences in which aniinal appears is at base of matrix column ) AD19d ■78 ■76 ■68 ■73 ■57 68 30 182 -84 -73 -59 -74 56 39 34 50 -52 -67 32 -66 -54 7 ADZe (■89) AD27 {•96) ADp99 AD07Q I 012 ■81 -83 ■64 ■51 48 51 ■29 ■24 •69 ■33 ■50 59 85 AooeQ AD24 cgo) ADZSCM) AD22C93-*) 007 022 024 009 001 003 010 Doe 023 004 Oil 019 026 DOS 025 006 021 002 012 Figure 5 Top: dendrogram illustrating intrapod groups and relationships among individuals with CAI values calculated at the population level. (Relationships among intrapod groups shown in Fig. 2.1 Middle: matrix showing CAI values between pairs of individuals calculated at the pod level for ADS pod. Numbers at the base of each column in the matrix are the number of valid photo sequences in which that individual appears. Bottom: inferred genealogical trees. and AN48; ANl 1 and calves AN47 and AN49). These females became most closely bonded to their own calves rather than to their mothers. One juvenile ^oiale, AN19, had a stronger bond with apparent ju- venile sibling AN18 (CAI=16) than with his appar- ent mother, reproductive female AN17 (CAI=4). The juvenile AB18 had a stronger bond (CAI=58) with a young calf AB41, in his intrapod group than with his apparent mother, reproductive female AB7 lCAI=22). However, in the cases of all five of these whales, the strongest bond with an adult female re- mained with their apparent mothers. 910 Fishery Bulletin 97(4), 1999 ID 5 20 30 40 50 60 70 80 90 100 D16 361 □20 22 1% D18 9 013 42 -50 ■M 361 AD16Q 017 24 -11 37 « AD14 014 45 50 •37 39 12 377 015 -17 ■20 -6 -23 -14 102 016 □20 018 013 017 014 015 AD18(? AD20C88) AD 13d AD15 AD17 AB1 ABl (■88) Whale Designaton Whale Deceased Known Date of Birth Known Relationship Proposed Relationship j photographic sequences J appears is at base of (Number o1 valk in wtiicti anima matrix column.) Figure 6 Top: dendrogram illustrating intrapod groups and relationships among individuals with CAI val- ues calculated at the population level. (Relation- ships among intrapod groups shown in Fig. 2.1 Middle: matrix showing CAI values between pairs of individuals calculated at the pod level for AD16 pod. Numbers at the base of each col- umn in the matrix are the number of valid photo sequences in which that individual appears. Bot- tom: inferred genealogical trees. Table 4 Cole's Association Index ( CAI ) values for male killer whales and thei r mothers. Ages estimated as described in meth- ods; age 5 over 20 years are mimimum ages. Estimated Estimated approx. CAI approx. CAI Whale age (yr) Value Whale age (yr) Value ABl 33 25 AJ9 18 32 AB3 30 20 AJ16 27 48 AB24 25 39 AJ17 29 18 AB35 19 35 AJ21 19 31 AJ25 22 48 AK4 27 29 AD3 27 50 AKl 27 20 AEl 32 25 AN3 32 31 AE3 18 21 AN 7 24 17 AE6 16 38 AN14 19 47 AE14 19 24 AN21 25 54 AE9 32 31 AN24 15 55 AJl 31 39 AN25 24 28 AJ2 27 21 AN30 16 54 AJ7 19 55 AN33 23 41 ANl 32 66 Of the 45 proposed maternal lineages (Figs. 3-12), 25 included two generations, 16 included three gen- erations, and four had four generations. In all but one of the four-generational trees, the oldest matriarch died during the study. We identified four matrilineal intrapod groups that appeared destined to die out. These were the single adult male, AB3 ( Fig. 4); the AB 10 subpod which consisted of the apparently postrepro- ductive female AB 10 and 3 adult male sons, AB4,AB5, and ABll (Fig. 6); the apparently postreproductive AJ12 and her adult male son, AJ16 (Fig. 9); and the apparently postreproductive AN 34 and her adult male son, AN21 (Fig. 10). Three of the remaining matrilin- eal groups had produced two reproductive females in one generation and were growing, and eight had pro- duced one reproductive female and were relatively stable. The fate of the other matrilineal groups will be determined later as offspring bom during the study mature and their reproductive potential is realized. Discussion The strength of bonds between male offspring and their mothers was highly variable (Table 4). Simple regression analysis indicated that the strength of bonds tended to diminish with age, but not signifi- cantly (Fj ^.= 1.62; P=0.215). Direct observations indicated that resident killer whales in Prince William Sound do not associate ran- domly with one another, but rather tend to associate with specific individuals. Statistical analysis of as- sociations in photographic sequences supports this Matkin et al : Association patterns of Oranus orca 911 20 30 40 50 60 70 80 90 100 8 lU S Ul E02 E20 E16 E01 E04 EOS E07 Ell E19 E18 Eoe E13 E12 E15 E10 EOS EOS E17 E06 E14 1587 77 64 24 15 16 12 -13 -41 -52 -18 -37 -73 -54 -38 -53 -59 -64 -61 -46 EOB AE07C? 149 40 36 • 27 • -74 -77 -78 * • • -90 -73 -50 -59 -77 -62 -60 E20 A^M? 1021 17 • 13 • -51 -55 -55 ♦ -30 • -54 -50 -59 -67 -65 -57 -« E16 837 25 129 21 38 8 8 -53 44 -55 • -54 -82 -54 -54 -95 -80 -86 -38 -47 -72 -16 • ■58 76 -61 • -55 -84 -48 E01 -68 E04 :oid AEd2g AE03C! 904 5 157 -60 -51 -77 -53 • -57 1 -56 -70 -61 • -43 -49 -62 -13 -81 -52 10 -56 • -46 9 -31 E03 86 607 16C5C89) ^E209(■95) AFiion 1923 59 57 45 46 -34 -to -19 -20 -40 -48 -45 -5 Ell 410 43 957 • 33 19 " A2 -44 -34 -34 -23 -23 -12 -42 -49 -49 -54 -52 28 E19 -14 E18 AE09d Afiog 48 • • • -14 -89 14 • 2 24 EOB AE15('88) 998 -9 -33 -29 -34 -56 -58 -63 0 E13 121 4 16 31 -69 . -6 -20 E12 AE089 868 46 11 -32 -41 -41 -12 E15 AE119 ~r- 1346 27 -45 -36 -44 1 3 E10 AE14c5 AE13C85) AE18c5C90) AE19dC93 792 -49 -59 -50 11 EOS AEOS? 1012 46 835 38 36 -28 EOS 1 1 -29 E17 AE06C5 AE179C89) 955 -17 £06 1033 E14 ABl Whale Designation AB1 Whale Deceased (■88) Known Date of Birth Known Relationship Proposed Relationship (Number of valid photographic sequences In which animal appears Is at base of matrix column ) Figure 7 Top: dendrogram illustrating intrapod groups and relationships among individuals with CAI values calcu- lated at the population level. (Relationships among intrapod groups shown in Fig. 2.1 Middle: matrix show- ing CAI values between pairs of individuals calculated at the pod level for AE pod. Numbers at the base of each column in the matrix are the number of valid photo sequences in which that individual appears. Bot- tom: inferred genealogical trees. observation. It is important to appreciate that in this analysis (Fig. 1 and subsequently Figs. 3-12), results were obtained by using a continuous clustering pro- cess in which all whales were treated as individuals 912 Fishery Bulletin 97(4), 1999 and were linked into progressively larger groups until they formed a single large group. This analysis ex- amines the relationships between individuals in the entire population. We postulate that the significant long-term asso- ciations observed were based on genealogical rela- tionships. Judged from known relationships (moth- ers and offspring born during the study), the strength of bonds among individuals within pods appeared directly correlated with their degree of relatedness. 20 30 40 50 60 70 80 90 100 D3 27 Ills •17 944 AIO39 ns -35 1054 102 101 •25 1005 AI01 d AI02C5 -29 -36 835 AlOSd AI06 d A1049 AID? cse) AB1 Whale Designation AB1 Whale Deceased (■88) Known Date of Birth Known Relationship Proposed Relationship (Numbei ol valic in which animal matrix column.) photographtc sequences appears is at base ol Figure 8 Top: dendro^am illustrating intrapod groups and relationships among individuals with CAI values calculated at the population level. (Rela- tionships among intrapod groups' shown in Figure 2.1 Middle: matrix showing CAI values between pairs of individuals calculated at the pod level for AI pod. Numbers at the base of each column in the matrix are the numbers of valid photo sequences in which that individual appears. Bottom: inferred genealogical trees. Our findings that resident killer whales in Prince William Sound are organized in statistically identi- fiable pods and intrapod groups are similar to re- sults from studies of resident killer whales in Brit- ish Columbia and Washington State (Heimlich- Boran, 1986; Bigg et al., 1990). Bigg et al. (1990) defined a pod as a group of individuals that traveled together at least 50% of the time. All the resident pods described in Prince William Sound fit that defi- nition. Pod membership is also supported by pod- specific vocal dialects in both areas (Bigg et al., 1990; Ford, 1991; Fordi). Pod delineations based on direct obser- vation were supported by association analysis in all but two cases. Both dis- crepancies appeared in the final stages of the agglomerative linking procedure. Although each pod was a distinct clus- ter, AD 16 pod was found to be linked to AK pod with a PCC value of eight, at somewhat higher than the level of link- age between other pods. This was not supported by direct observation and was apparently an artifact of small sample size. AD 16 pod was infrequently photo- graphed and was often part of multipod groups that included AK pod. The ABIO subpod was linked with AI pod (PCC=4) before being linked with other members of AB pod. AI pod fre- quently traveled with AB pod early in the study, and we suspect that AI pod (7 whales in 1996 ) was in the final stages of a gi-adual split from the then 35-member AB pod when the study began in 1984. AI pod traveled more independently of AB pod in later years. The pod-specific dia- lects for AI and AB pods are very similar (Ford^), supporting this hypothesis. The preponderance of adult males in AI pod (4 out of 7 whales in 1996) may have contrib- uted to the independence of this matrilin- eal group. Bigg et al. (1990) found that matrilineal groups with a high percentage of males tend to travel more independently. This was also evident for the ABIO subpod, in which 3 out of 4 members were adult males. ABIO subpod often traveled a dis- tance away fi-om the remainder of AB pod. Direct observation and photographic analysis also indicated that some males 108 04 D6 05 loe 01 Ford, J. K. B. 1997. Vancouver Public Aquarium, P.O. Box 3232, Vancouver, B.B., Canada. Personal commun. Matkin et a\: Association patterns of Oranus orca 913 20. 30 40 50 60 70 80 90 100 « ^ h J3B jm jit m jx SH m J3* AJ089 AJ<52C5 AJ039 AJ049 AJ10 AJ29 ('89) AJ35('93) ■AJ36C93) AJ30 (W) AJ34 caa) AJ239 AJ249 AJ25d AJ229 AJ059 AJOld AJ069 AJIld- AJ149 AJZSCSS) AJ31 ('92) AJ26 CSS) AB1 AB1 (■88) Whale Designation Whale Deceased Known Data of Birth Known Relationship Proposed Relationship i photographic sequences appears is at tMse of in which anima malrtx column.) AJ32 ('92) AJ129 AJied AJ139 AJISd AJ33 (■92) AJ39 Cge) AJ209 1_ Aji/d AJ2ld AJ27('87) /U38 ('94) AJ189 AJ09d AJ19d AJ37 C^A) Figure 9 Top: dendrogram illustrating intrapod groups and relationships among individuals with CAI values calculated at the population level. (Relationships among intrapod groups shown in Fig. 2.) Middle: matrix showing CAI values between pairs of individuals calculated at the pod level for AJ pod. Numbers at the base of each column in the matrix are the number of valid photo sequences in which that individual appears. Bottom: inferred genealogical trees. 914 Fishery Bulletin 97(4), 1999 AKOld 20 30 40 50 60 70 80 90 100 K02 K13 KIO K09 K05 7: 75 54 61 M5 51 47 36 470 KM 2V 21 13 21 21 667 K03 17 .34 12 7 23 16 44S K06 -81 .1)3 -94 ■86 -65 -75 -58 944 AK069 Kll -82 -«7 -85 •74 46 308 K08 -K2 -92 -85 -86 -66 -60 -66 34 18 586 AK07O I AK12 ('93)9 AK08 CSS) ABl Whal« Designation AB1 Whale Deceased (■88) Known Date of Birth Proposed Relationship (Number of valK in which anima matrix column) photographic sequences appears is at base of AK11 ^■9^) AK03ase of ) A 29 r96i Ni09 AN 359 AN 119 AN •95) AN51 [ AN40 1 r D("96) AN47 (Number of v In wtilcti anir matrix column (•« A 1) N4j ANSI '92) AN49 ( 1 AN38C5C87) 1 " AN46 ('92) t \9lf Fic |ure 11 Top: dendrogram illustrating intrapod groups and relationshi ps among individi lals with CAI values calculated at the ] m )ulation level. (Relationships among intrapod groups shown in Fig. 2 ) Middle: matrix t bowing CAI values between pairs of ii id viduals calculated at the po d level for ANIO pod. Numbers at the b ase of each co un in in the matrix are the number of v alid photo | sequences in which that individual appears. Bottom: inferrec genea logical tree s. 916 Fishery Bulletin 97(4), 1999 from their pods. Through association analysis, we expanded this definition to include groups that travel in unison with their pod but rarely mix with other groups within the pod. This latter situation occurred only for AB pod, which was divided into three subpods. By direct observation and examination of 20 30 40 50 60 70 80 90 100 z z H\3 N1S N14 PC4 fCI N1E KS NU HSl Kit WB Ha KB POl NM N19 NZ3 KB NS rcE M2 m n 58 «8 II 1 » « .61 ^» NM KQ «9 NO AN 299 AN24d AN43 ('90) AN 139 AN289 AN30d AN349 AN21(5 AN169 ANi79 ANi4C5 ANSad ^''^ AN^9d AN18 AN37 ('87) AN279 AB1 Whale Designation AB1 Whaie Oecaasod (■88) Known Data o( Birth Known Relationahip Proposed Relationship (Number ot vaiW In ¥v^lch animai matJix column.) photographic sequences appears is at tjase of AN04d AN239 L_ AN32 AN269 AN2Sd ANSS CBS) I AN42('90) AN209 AN319 AN22 I AN44C90) AN39('88) let KB « 86 n n « 2 52 52 71 10 Figure 12 Top: dendrogram illu.strating intrapod groups and relationships among individuals with CAI values calculated at the population level. (Relationships among intrapod groups shown in Fig. 2.) Middle: matrix showing CAI values between pairs of individuals calculated at the pod level for AN20 pod. Numbers at the base of each column in the matrix are the number of valid photo sequences in which that individual appears. Bottom: inferred genealogical trees. Matkin et al : Association patterns of Oranus orca 917 photogi'aphs of this pod, it was determined that AB pod nearly always traveled as a unit; however, in the dendrogram linking intrapod groups (Fig. 2) it ap- peared as three separate groups (PCC>0). This find- ing indicated that although they were traveling to- gether, the subpods tended not to mix with the rest of the pod. Communities were described by Bigg et al. (1990) as closed populations of pods that associated with one another. They described two geographically dis- tinct communities of resident killer whales (north- ern and southern residents) that had separation in range near the middle of Vancouver Island, British Columbia. We found no separation of pods into com- munities by this definition in our area, although our study discerned matrilineal (intrapod) groups, subpods, and pods. Resident whales from AF and AG pods photographed regularly in southeastern Alaska were observed swimming with the pods described in this study (Matkin et al., 1997). One of the pods de- scribed in this paper, AD5 pod, was photographed off Kodiak Island. It thus appears that geographic boundaries do not delineate communities for resident killer whales from southeastern Alaska to Kodiak Island. The association analysis for individuals within pods strongly suggested that strength of bond was directly correlated to degree of relatedness. There was only one case in which statistical analysis indi- cated that offspring born during the study did not maintain the strongest bond with its mother. The mother AB8 and her sibling AB18 both died at the time of the Exxon Valdez oil spill. AB8 left her year- old offspring. AB41 (born 1988). Association analy- sis indicated that AB41 was more closely linked (CAI=58) to its mother's apparent sibling, AB18, than to its mother, AB8 (CAI=50); the young whale AB41 died several years later ( 1993-4). Five whales that were juveniles at the beginning of the study were most closely associated with whales other than their mothers. Three were females, AK07, AN08, and ANll, each of which produced offspring during the study. Following these births, they were more closely bonded to their offspring rather than to their own mothers. This finding demonstrates that new mothers, when they produce calves, develop dis- tance from their own mothers (Table 4). The tendency of females with offspring to travel farther from their mothers than the distance prior to first reproduc- tion suggests a process of separation that may also be basic to new pod formation. Our proposed genealogical trees suggest that jntrapod groups are matrilineal groupings of moth- ers and their descendants (Figs. 1 and 3-12). There was no immigration or emigration of male or female offspring from these natal groups over the course of our study. These extremely stable matrilineal groups appear to be the foundation of resident pod social structure in Prince William Sound. This is similar to results from studies in British Columbia and Wash- ington State (Bigg et al., 1990). We were most confident in genealogical trees for pods that were most frequently photographed, such as AE and AK pods, and less confident in trees for the much less frequently observed ADS and AD16 pods. The large number of mortalities in AB pod also made construction of genealogical trees more diffi- cult. The greatest potential source of error in genea- logical assignment was the death of the mother of a young whale prior to the study, in which case the young whale would likely travel with its closest fe- male relative. However, since the annual natural mortality rate for reproductive females is extremely low (0.0048 according to Olesiuk et al., 1990), this source of error was probably insignificant except fol- lowing disasters such as the high loss of reproduc- tive females after the Exxon Valdez Oil Spill (Matkin etal., 1994). In all but one of the resident pods we examined, the total number of whales increased over the pe- riod of the study, indicating that a majority of matri- lineal groups have been growing or dividing (or do- ing both) over the past decades. The exception was AB pod, which declined during this period from 35 to 23 whales (28 deaths; 16 births). Six of the mor- talities occurred during 1985 and 1986 when there was a conflict with the sablefish (Anaplopoma fim- bria) fishery (Matkin et al., 1994). During this pe- riod apparent bullet wounds were observed in 16 whales. Fourteen of the mortalities in AB pod occurred in the year and a half following the 1989 Exxon Valdez oil spill (Matkin et al., 1994). Some of the matrilin- eal groups in the pod are nearly extinct as a result of these deaths. An adult male, AB03, appears to be the last member of a once-large matrilineal group linked by the apparent sisters AB06 and AB07 (Fig. 1; AB03 has since died). Another large matrilineal group ( matriarch AB09) has been reduced to a single orphaned 5-year-old, AB45. Many of the mortalities have been those of juveniles (13) or reproductive fe- males (4) and have severely reduced the reproduc- tive potential of the pod. The statistical support for the existence of intrapod groups and the apparent lack of emigration or immi- gration into these groups allows for determination of demographic changes, both natural and induced, within resident killer whale populations. The de- tailed delineation of social structure in resident killer whales provides a unique opportunity for monitor- 918 Fishery Bulletin 97(4), 1999 ing and modeling of these populations. This situa- tion appears to be unique among social mammals (e.g. the lack of male dispersal), with the possible exception of the long-finned pilot whale. In that spe- cies, molecular typing revealed that pod members also form extended family groupings from which in- dividuals do not disperse (Amos et al., 1993). Mo- lecular typing of individual whales described in this study is currently in progress (Barrett-Lennard-). Acknowledgments Work from 1989-92 was funded as part of the dam- age-assessment program of the Exxon Valdez Oil Spill (EVOS) Trustee Council. The EVOS Trustee Council also provided funds in 1995 for field work and analysis as part of the oil-spill restoration pro- gram. Work in 1984 was supported by Hubbs Sea World Research Institute, and fieldwork in 1986 was supported by Alaska Sea Grant and the National Marine Mammal Laboratory. Data for killer whale pods in Prince William Sound in 1985, 1987, 1988, 1993, and 1994, and in March 1989 following the EVOS, were provided by the North Gulf Oceanic So- ciety. The Society was funded by private donations, the Alaska State Legislature, the Sail Alaska Foun- dation, and the Alaska Sea Grant program. The project would not have been possible without the participation of L. Barrett-Lennard, K. Heise, O. von Ziegesar, and K. Balcomb-Bartok, who led field efforts at various times. Field assistance or other substantial contributions were made by R. Angliss, L. Daniel, K. Englund, F. Felleman, M. Freeman, B. Goodwin, M. Hare, M. James, L. Larsen, J. Lyle, E. Miller, L. Saville, C. Schneider, R. Blancato, S. A. Sikema, K. Turco, and E. Weintraub. E. Miles of Miles Photo Lab provided all photographic development and printing services. Literature cited Amos, B., C. Schlotterer, and Diethard Tautz. 1993. Social structure of pilot whale.s revealed by analyti- cal DNA profiling. Science (Wa.sli. D.C.i 260:670-672. Balcomb, K. C, J. R. Boran, and S. L. Heimlich. 1982. Killer whales in Greater Puget Sound Rep Int Whal. Comm. 32:681-686. Bigg, M. A. 1982. An as.ses.sment of killer whale UJrnniis oria i stocks off Vancouver Island, British 'Columbia. Rep. Int. Whal. Comm. .32:6.5.5-666. 2 Barrett-Lennard. L. 1998. Dep. Zoology, Univ. British Colum- bia, Vancouver. B.C.. Canada. Personal commun. Bigg, M. A., G. M. Ellis, and K. C. Balcomb III. 1986. The photographic identification of individual cetaceans. Whalewatcher: .J. Am. Cetacean Soc. 20:10-12. Bigg, M. A., G. M. Ellis, J. K. B. Ford, and K. C. Balcomb III. 1987. Killer whales: a study of their identification, genealogy and natural history in British Columbia and Washington State. Phantom Press, Nanaimo, British Columbia. 79 p. Bigg, M. A., P. F. Olesiuk, G. M. Ellis, J. K. B. Ford, and K. C. Balcomb III. 1990. Social organization and genealogy of resident killer whales iOrcinus orca) in the coastal waters of British Co- lumbia and Washington State. Rep. Int. Whal. Comm. Spec. Issue 12:386-406. Cole, L. C. 1949. The measurement of interspecific association. Ecol. 30:411-24. Ellis, G. M. 1987. Killer whales of Prince William Sound and South- east Alaska: a catalogue of individuals photoidentified, 1976-1986. Hubbs Sea World Res. Inst. Tech. Rep. 87- 200, 76 p. Ford, J. K. B. 1991. Vocal traditions among resident killer whales iOrcinusorca) in coastal waters of British Columbia. Can. J. Zool. 69:1454-1483. Ford, J. K. B., G. M. Ellis, L. G. Barrett-Lennard, A. B. Morton, R. S. Palm, and K. C. Balcomb. 1998. Dietary specialization in two sympatnc populations of killer whales tOrcinus orca) in coastal British Columbia and adjacent waters. Can. J. Zool. 76(8):1456-1471. Heimlich-Boran, S. L. 1986. Cohesive relationships among Puget Sound killer whales. In B. C. Kirkevold and J. S. Lockard (eds.), Behav- ioral biology of killer whales, p. 251-284. Alan R. Liss, New York, NY. Heise, K., G. Ellis, and C. Matkin. 1992. A catalogue of Prince William Sound killer whales, 1991. North Gulf Oceanic Society Homer, AL. 51 p. Hoezel, A. R., Dahlheim, M., and Stern, S. J. 1998. Low genetic variation among killer whales iOrcmus orca ) in the eastern North Pacific and genetic differentia- tion between foraging specialists. Heredity 89:121-128. Hoezel, A. R., and G. A. Dover. 1991. Genetic differentiation between sympatric killer whale populations. Heredity 66:191-195. Johnson, S. C. 1967. Hierarchical clustering schemes. Psvchometrika 32:241-54. Leatherwood, S., K. C. Balcomb HI, C. O. Matkin, and G. Ellis. 1984. Killer whales [Orciniin urea) in southern Alaska. Hubbs Seaworld Res. Inst. Tech. Rep. 84-175. 54 p. Leatherwood, S., C. O. Matkin, J. D. Hall, and G. M. Ellis. 1990. Killer whales, Orc;>7;;,s orca, photo-identified in Prince William Sound, Alaska, 1976 through 1987. Can. Field- Nat. 104(31:362-371. Matkin. C. O., G. Ellis, M. Dahlheim, and J. Zeh. 1994. Status of killer whales in Prince William Sound, 1984-1992. In T. Loughlin (ed.). Marine mammals and the Exxon Valdez, p. 141-162. Academic Press, San Di- ego, CA. Matkin, C. O., D. Matkin, G. M. Ellis, E. Saulitis, and D. McSweeney. 1997. Movements of resident killer whales between south- eastern Alaska and Prince William Sound, Alaska. Mar. Mamm. .Sci. 13i3i:469-475. Matkin et al : Association patterns of Orcinus orca 919 Morton, A. B. 1990. A quantitative comparison of the behavior of resident and transient forms of killer whale off the central British Columbia coast. Rep. Int. Whal. Comm. Spec. Issue 12:245-248. Olesiuk. P. F., M. A. Bigg, and G. M. Ellis. 1990. Life history and population dynamics of resident killer whales (Orcinus orca) in the coastal waters of Brit- ish Columbia and Washington State. Rep. Int. Whal. Comm. Spec. Issue 12:209-244. Saulitis, E. 1993. The behavior and vocalizations of the AT group of tran- sient killer whales in Prince William Sound, Alaska. Master's thesis, Univ. Alaska, Fairbanks, AK, 113 p. Saulitis, E. L., C. O. Matkin, L. Barrett-Lennard, K. Heise, and G. Ellis. In press. Foraging behavior of transient killer whales (Orcinus orca) in Prince William Sound, Alaska. Mar. Mamm. Sci. Stevens, T. A., D. A. Duffield, E. D. Asper, K. G. Hewlett, A. Bolz, L. J. Gage, and G. D. Bossart. 1989. Preliminary findings of restriction fragment differ- ences in mitochondrial DNA among killer whales (Orcinus orca ). Can. J. Zool. 67:2,592-2,595. 920 Abstract.— We studied fine-scale ver- tical distribution of northern anchovy, EngrauliR mordax, eggs and larvae and larvae of associated species at two sta- tions off southern California in March- April 1980 using a Manta ( neuston ) net and a MESSHAI (multiple opening- closing net I sampler. A pump and fluo- rometry system was used to obtain chlo- rophyll profiles. A storm with associ- ated >feeavy seas interrupted sampling for two days at the inshore station and provided an opportunity to compare pre- and poststorm egg and larval dis- tribution and abundance. A total of 95,552 fish larvae were taken in the MESSHAI (63 tows) and Manta nets (41 tows), representing 49 taxa (genera or species) in 27 families. Engraulis mordax was the most abundant fish ( 95"f of the total), followed by Leuroglossus stilhius. Genyonemus lineatus, Steno- brachius leucopsarus, Sebastes spp., Seriphiis politus. Peprilus simillimus, Paralichthys californicus, Citharichthys spp., and Merluccius productus. Anchovy eggs and larvae had a shallow distribu- tion: 90'f of larvae and 95'7f of eggs were found m the upper .30 m. Peak egg den- sity was found in the neuston; peak lar- val density, in the 10-20 m stratum. Larvae of shallow-living shelf species (G. lineatus, P. simiUimus.P.califoniicus) typically occurred in the upper 20-30 m, whereas larvae of predominantly deeper- living demersal species (Sebastes spp., M. productus I were found in the upper 80 m. Midwater species (L. stilbius. S. leucopsarus) occurred at least to 200 m. Anchovy eggs decreased in number af- ter the storm. Anchovy larvae declined even more sharply, despite an increase in zooplankton and potential larval fish prey, suggesting that starvation may have resulted from disturbance of microscale food patches. Alternatively, larvae may have been advected away from the site. Shallow-water shelf spe- cies were rare at the inshore station before the storm but appeared suddenly afterwards as a result of seaward advec- tion of shelf water by Ekman transport. Vertical distribution of eggs and larvae of northern anchovy, Engraulis mordax, and of the larvae of associated fishes at two sites in the Southern California Bight H. Geoffrey Moser Southwest Fisheries Science Center PO Box 271 La Jolla, California 92038 E-mail address gmoseraiucsd edu Tilman Pommeranz Gotennng 66, D-53913 Swisttal, Odendorf Germany Manuscript accepted 26 October 1998. Fish Bull. 97: 920-943 ( 1999). The use of ichthyoplankton tech- niques for estimating the biomass of fish stocks has increased mark- edly over the past two decades (Hunter and Lo, 1993). Information on the horizontal and vertical dis- tribution of eggs or larvae of a tar- get species is a requirement for quantitative sampling of these stages and for interpreting data from integrated net tows and from recently developed samplers such as the continuous underway egg pump (Checkley et al., 1997). The daily egg production method ( DEPM ) was developed at the Southwest Fisher- ies Science Center to provide annual biomass estimates needed to man- age northern anchovy, Engraulis mordax, an important coastal pe- lagic species inhabiting the Califor- nia Current region (Lasker, 1985). During the development of this method a cruise was conducted off southern California that employed a MESSHAI, a sampler with open- ing-closing nets and environmental sensors, to determine fine-scale ver- tical distribution of eggs and larvae of £. mordax (see preliminary data in Pommeranz and Moser, 1987). This information was needed to supplement Ahlstrom's (1959) ver- tical distribution study based on 22 Leavett net tows taken on 9 cruises over a 14-year period off southern California and Baja California. The MESSHAI study supplied informa- tion on depth, ambient tempera- ture, and diurnal periodicity of E. mordax eggs needed to develop the DEPM (Lasker, 1985). To compare distributions at different bottom depths, sampling was conducted at two localities: an offshore site over the Santa Catalina Basin and an inshore site over the upper conti- nental slope off Dana Point, Cali- fornia. Anchovy larvae, the domi- nant species in the MESSHAI samples, were counted and mea- sured for subsequent growth and mortality studies. A storm inter- rupted sampling at the inshore sta- tion and provided an opportunity to study the influence of water column disturbance on the abundance and depth distribution of eggs and lar- vae of anchovy and associated spe- cies. Mullin et al. (1985) described the effects of this storm on the abun- dance and vertical structure of phy- toplankton, zooplankton, and total fish larvae on the basis of pump samples taken at the same time as the MESSAI sampling at a locality I Moser and Pommeranz: Distribution of eggs and larvae of Engraulis mordax 921 just shoreward of the inshore MESSHAI site. This paper presents a pre- hminary analysis of the fine- scale vertical distribution of anchovy eggs and lai-vae and the larvae of other taxa, with comparisons of the offshore and inshore sites and of the prestorm and poststorm peri- ods at the inshore site. A sepa- rate paper addressing age-spe- cific vertical distribution and mortality of anchovy eggs and larvae is in preparation. Methods and materials 34 00 33 30 33°00' Cruise 8003-EB Stations 27 March, 1980 29 March-6 April, 1980 Cruise 8003-EB was conducted aboard RV Ellen B. Scripps at two sampling sites in the Southern California Bight (SCB; Fig. 1). The first site 32°3o' (33°11.1'N, 118°16.5'W), occu- pied 19-27 March 1980, was located 13 km south of the east end of Santa Catalina Island, over an -1150 m bottom depth. This site lies over the Santa Catalina Basin, one of 13 deep basins that charac- terize the offshore region of southern California (Ebeling et al., 1970; Eppley 1986). The midwater fish fauna of these basins is a mix of subarctic, tran- sitional species of the California Current and spe- cies from the eastern tropical Pacific and central water masses whose distributions extend into the region. The site is close enough to Santa Catalina Island to have larvae of shelf species (Hobson and Chess, 19761 in addition to larvae of slope and basin floor species ( Cross, 1987 ). The second site ( 33"28.5'N, 117°47.0'W), occupied 29 March-6 April 1980, was -4.5 km off Dana Point, over the upper continental slope at approximately 430 m bottom depth. Here the adjacent shelf narrows to a minimum of about 3 km but supports rich fish communities associated with rocky reefs and kelp forests (Feder et al., 1974; Cross and Allen, 1993) and soft-bottom habitats (Meams, 1979; Love et al., 1986; Allen and Herbinson, 1991; Cross and Allen, 1993). The continental slope, also relatively narrow here, supports a fish assem- blage typical of other southern California slope re- ^gions (Cross, 1987). The first site corresponds to Cali- fornia Cooperative Oceanic Fisheries Investigations (CalCOFI) station 90.2, 35.8 and the second to sta- ll?" San Diego Figure 1 The Southern California Bight, showing stations occupied on Cruise 8003-EB (modi- fied from Shepard and Emery, 1941). tion 90.0, 28.2. Sampling at the offshore station was interrupted twice by heavy seas (0055, 21 March to 0908, 23 March; and 1730, 26 March to 0836, 27 March ). A storm with associated heavy seas curtailed sampling at the inshore site at 0315, 1 April; the sta- tion was reoccupied at 0655, 3 April. The principal samplers used were the Manta net (Brown and Cheng, 1981) for sampling the surface layer (0-15.5 cm) and the MESSHAI for sampling discrete depth strata in the upper 200 m. The MESSHAI sampler was a modified Gulf-V sampler (Arnold, 1959; Nellen and Hempel, 1969) with a 25x25 cm mouth opening, six opening and closing nets operated from a computer deck unit, and an environmental sensing package (Pommeranz and Moser, 1987). All nets were fitted with 300-^m mesh. A sampling sequence for a day or night station con- sisted of a five-minute Manta tow, followed by an oblique MESSHAI tow that sampled 10-m strata from 50 m to the surface, an oblique MESSHAI tow which sampled 40-m strata from 200 m to the sur- face, a second MESSHAI tow which sampled 10-m strata from 50 m to the surface, and a second Manta tow. A total of 17 Manta tows ( 10 day, 7 night) were taken at the offshore station, 24 (13 day, 11 night) at 922 Fishery Bulletin 97(4), 1999 the inshore station (Figs. 2 and 3). All MESSHAI tows were stepped oblique hauls. In the shallow tows the sampler was lowered to 50 m and retrieved through five 10-m strata, each in 2.5-m steps lasting two min- utes, with a total of five horizontal phases per stra- tum. Depth, temperature, and flowmeter profiles were recorded by the deck unit throughout each tow. I-Mar 19^Mar 20H hMar 23-4-Mar 24-|-Mar 25H hMar 27^ » 30 ^Mar 19 + Mar 20^ hMar 23+Mar 24-|-Mar 25H hMar 27 H 160 - 200 - Figure 2 Diagram showing chronology of MESSHAI tows, depth strata sampled, and isotherm.s for the offshore station ifrom Pommeranz and Moser, 1987). i Above i Shallow (50 m) and Manta tows; the dots represent average depths for each net; the seven complete vertical lines show when deep i200 m) tows were taken. Tow numbers for Manta tows are" given above solid triangles and tow numbers for MESSHAI tows are given at the bottom of each series of vertical lines. Iso- therms are shown in 0.5 C intervals. (Below) Deep tows; the 17 lines extend- ing down to 50 m indicate when shallow tows were taken. Isotherms are shown in 1 C intervals. Bucket surface temperatures were taken periodically to calibrate the MESSHAI temperature sensor. In the deep tows the sampler was lowered to 200 m and retrieved through five 40-m strata, each in five steps lasting two minutes. The total number of successful MESSHAI tows was 24 (shallow: 9 day and 8 night; deep; 4 day and 3 night) at the offshore station and 36 (shallow: 12 day and 12 night; deep: 6 day and 6 night) at the in- shore station. Improper preserva- tion or breakage resulted in the loss of anchovy eggs in net 4 (20-30 m stratum) of tow 42 and the entire sample from net 3 (30-40 m stra- tum) of tow 61. Three MESSHAI tows were unsuccessful owing to technical difficulties, and the samples were subsequently dis- carded. All tows were made at a ship speed of -1.5 knots. Before each net tow series at the inshore station, a pump and fluorometry system (Lasker, 1978) was used to obtain a chlorophyll profile down to 40 m depth. Water samples were taken at the surface and at the chlorophyll maximum for chlorophyll-a and phaeophytin extraction and fluoro- metry, and for identification and counts of major phytoplankton taxa (analyses not included in this pa- per). Oblique bongo net samples were taken at each station at a ship speed of about 1.5 knots; except for length-frequency information, data from these tows are not addressed in this paper. Most samples were preserved in S'^ buffered formalin; selected samples were preserved in 80% ethyl alcohol for analysis of otolith daily growth rings in anchovy lar- vae. Prior to sorting, wet zooplank- ton displacement volume of each sample was measured using stan- dard techniques (Kramer et al., 1972). Anchovy larvae and eggs were identified and removed during the plankton sorting process; eggs were classified by stage according to the criteria of Moser and Ahlstrom ( 1985 ) and length measurements of larvae were taken to the nearest 0.5 mm. Larvae of other fish were identified to the lowest taxon possible Moser and Pommeranz: Distribution of eggs and larvae of Engraulis mordax 923 rMar 29- NIGHT Manta-tow *18 19 2021 22 and body lengths were measured to 0.1 mm. We used analysis of variance (AN OVA) to compare the pre- and poststorm densities of anchovy eggs and larvae for shallow strata and also for deep strata at the inshore and offshore stations. Day and night data were pooled for each of these categories because the data did not indicate a diurnal shift in vertical distribution of eggs or larvae. All data were log-transformed for sta- tistical analyses because the origi- nal data were highly skewed. Data from the shallow strata included the Manta tows, the MESSHAI shallow tows (0-50 m), and the upper 40 m from the MESSHAI deep tows. Data for the deep strata were from below 40 m from the deep MESSHAI tows. A two-way AN OVA was performed for the shallow strata, the two factors being station (inshore and offshore) and tow type (Manta, MESSHAI shallow, and MESSHAI deep). Inter- action terms were included in the analysis. One-way ANOVA was per- formed for data from the deep strata to compare anchovy larval density at inshore and offshore stations. In the analysis of prestorm and poststorm conditions two separate ANOVAs were performed because an initial ANOVA of anchovy eggs from the shallow strata indicated a strong interaction between tow type and pre- or poststorm period; one analysis was based on data from shallow and deep MESSHAI tows and the other was based on Manta tows only. A single two-way ANOVA test was performed for anchovy lar- vae because the density of anchovy larvae in Manta tows was low and the analysis adequately described the effect of the storm on larval density based on both shallow and deep MESSHAI tows. Results Environmental features The thermal structure of the study area was typical for March-April in that region of the SCB, with a 29 30c) 32n Figure 3 Diagram showing the chronology of MESSHAI tows, depth strata sampled, and isotherms at the inshore station. Symbols and labels are the same as those in Figure 2 (from Pommeranz and Moser. 19S7i. pool of relatively warm water subject to intrusions of advected upwelling plumes from the Point Con- ception area (Lasker et al., 1981; Lynn et al., 1982). The mixed layer was about 30-40 m deep at the off- shore station (Fig. 2) and only about 10-30 m deep at the inshore station (Fig. 3). Overall, the water was colder at the inshore station, which lacked the promi- nent 16.0^ isotherm present offshore. Average tem- perature was higher at the offshore station in all depth strata except 160-200 m (Fig. 4, A and B). At the inshore station after the storm the thermocline shoaled to 15-20 m and water temperature was lower 924 Fishery Bulletin 97(4), 1999 than before the storm at all depths below 80 m (Figs. 2; 3; 4, C and D). Wind and sea-state observations taken during the cruise and supplemented by data from the Naval Air Station at San Clemente Island, southwest of the offshore station, and the San Onofre Nuclear Gener- ating Station, just southeast of the inshore study area (Fig. 1), indicated an oscillating land-sea system whigh was more evident at San Onofre than at San Clemente (Fig. 5). During the night and morning, winds typically were from the NNE, after which they shifted through the south to the NW until late after- noon or evening. Average wind speed was about 10 knots at the offshore station and about 6 knots at the inshore station; average swell height was about 9 10 11 12 13 14 15 16 _l I I I I I I 120-160- 1 12 13 14 1 1 I 0-10- 16 _L_ 17 -J Inshore 10 11 12 13 14 11 12 13 14 15 16 I I I I I Post-Storm Pre-storm Temperature (C") Figure 4 Average temperatures for strata samplfd with the MESSHAI, based on data from the ME.SSHAI temperature sensor lAi Deep tows at inshore (dotsi and ofTshore (circle.s) stations; (B) shallow tows at inshore (dots) and offshore (circles) stations; (C) deep tows at inshore station before the storm (dots) and after the storm (circles); (Dl shallow tows at inshore station before the storm (dots) and after the storm (circles). 1.7 m and about 1 m at the two sites, respectively (Fig. 6). At the inshore station wind speed was more than 24 knots during the storm. Prior to the storm at the inshore station there was a moderate to strong chlorophyll maximum at 18-29 m (Fig. 7). Following the storm, the chlorophyll peak became shallower and generally broader. The peak was at 15 m on the morning of 3 April; by the evening of 4 April, the maximum was near the surface. On the following two days the peak reappeared and be- gan to deepen and strengthen. Average plankton vol- umes of Manta and MESSHAI samples from the upper 40 m were distinctly higher at the inshore sta- tion than at the offshore station (Fig. 8, A and B). Average volumes were generally higher in strata within the upper 40 m in poststorm samples compared with those taken prior to the storm (Fig. 8, C and E). Average plankton volume increased markedly in the 40-80 m stratum, whereas there was little change in deeper strata (Fig. 8E). Engraulis mordax eggs A total of 67,157 E. mordax eggs were collected in the MESSHAI and Manta nets. Anchovy eggs had a slightly shallower distribution than larvae, with approximately 95% of eggs in the upper 30 m ( Fig. 9, A and Bl. In shallow strata (including Manta, shallow MESSHAI, and the 0-40 m stratum of the deep MES- SHAI tows), average density was greater at the offshore station than at the inshore station (F-^ gr,=4.72, P=0.003 ). Yet egg density in the deep strata was similar at the two stations. Average egg density in the surface layer was more than double that in the 0-10 m stratum (Fig. 9B). In the shallow strata at the inshore station, egg densities in MESSHAI tows were reduced after the storm (Fj 32=12.73, P=0.001), whereas egg densities did not change much in Manta tows (F, ,,^=1.58, P=0.048) and in deep (>40 F,,,„=2.67, P= m) MESSHAI tows 0.132) (Fig. 9, C and F). Fish larvae A total of 95,552 fish larvae were taken in the MESSHAI and Manta nets, Moser and Pommeranz: Distribution of eggs and larvae of Engraulis mordax 925 \A^:^' ^^'^ V4-^'''V'^-^'Sf ^fe.^^^ — San Onofre San Clemente Island Offshore station Maf 1 9 — ' — Mar 20 — ' — Mar 2 1 L\///// ji r 22 1 Mar 23 — ' — Mar 24 Mar 25 — ' Mar 26 — ■ — Mar 27 - ^^/,. /7£,\l/£yM'///^yj/l ./i. /. San Onofre San Clemente Island _ Inshore station Mar 29 • Mai 30 '- Mar 3 Apr 2 ' Apr 3 ^ Apr 4 ' ^Apr5 1 Apr 6 " Figure 5 Wind direction and speed during Cruise 8003-EB based on data from San Onofre Nuclear Generating Station, the Naval Air Station on San Clemente Island, and obsen'ations made aboard RV£. B. Scnppa (from Pommeranz and Moser, 1987). representing 49 taxa (genera or species) in 27 families; 90,402 (957^ of the to- tal) were northern anchovy, Engraulis mordax (Table 1 ). Anchovy larvae were more than 40 times as abundant as the next most abundant species, California smoothtongue iLeuroglossus stilhius). White croaker (Genyonetnus lineatus) ranked third, and the northern lamp- fish iStenobrochius leucopsarus) ranked fourth, followed by the rockfish genus Sebastes. These five taxa consti- tuted 99'7r of the total larvae. Follow- ing in abundance were queenfish iSer- iphus politus). Pacific pompano (Pep- riltis simillinms), California halibut (Paralichthys californicus), the sand- dab genus Citharichthys, and Pacific _hake (Merluccius productits) (Table 1). Most of the larvae captured in the MESSHAI nets were at yolksac and preflexion stages. Relatively or in Fe.l 4 w Offshore station Mar 19 1 Mar 20 Mar 23- 4 Mar 24 * Mar 25 I ( Mar 27 Inshore station ! Mar 29 I Mar 30 < Mar 31 Apr 3 - Apr 4 ) Apr 5 — Apr 6 Figure 6 Swell direction and height during Cruise 8003-EB from observations made aboard RV E. B. Scnpps (from Pommeranz and Moser, 1986). fewer anchovy larvae <3.0 mm and relatively more anchovy larvae >3.0 mm were taken by the bongo 926 Fishery Bulletin 97(4), 1999 Temperature (°C) ■B 20 ~l — I — 1 — I — r 0 2 4 6 8 1012 0 2 4 6 8 1012141618 Chlorophyll-a ((ig/l) Figure 7 Chlorophyll-a l^ig/l) and temperature i C) profiles at the inshore station on Cruise 8003-EB. Tow numbers are given for shallow MESSHAI tows from which tempera- ture profiles were derived. net than in the MESSHAI tows (Fig. 10, A and B). The length-frequency distribution of larvae captured by the Manta net was bimodal, with peaks at 3.0 and 7.0 mm ( Fig. IOC ). For species other than north- ern anchovy, length distributions generally were similar for larvae caught in Manta and MESSHAI nets. Exceptions were a sample of 38 Sebastes spp. larvae captured in Manta net tow 18 (length range 5.0-9.4 mm, average length 7.2 mm) and an 8.0 mm presettlement Paralichthys californicits larva cap- tured in Manta tow 31. Engraulis mordax larvae Anchovy larvae had a shallow distribution, with approximately 90'7( of the larvae in the upper 30 m ( Fig. 1 1, A and B ). In shal- low strata, average larval density was greater at the Moser and Pommeranz: Distribution of eggs and larvae of Engraulis mordax 927 CO E o o o 1 E "-" 40-80 i= 80-120 O) 120-160 Q G Offshore I Inshore — r— 50 —I — 100 —I — 150 29 32 35 38 41 44 47 50 53 56 59 62 —I 250 Plankton volume (ml/1000m3) 500 T I I I I II I II I I 28 31 34 37 40 43 46 49 52 55 58 61 30 33 36 39 42 45 48 51 54 57 60 63 Tow number _ 0^0 40-80 E ra 80-120 GO SI g- 120-160 Q .^S\'>S\SH :;;.;.-.>.-'V.-.>.-'-.-.;.v;:'.-.V.-/.-.V..:-:t ,^.S^^;-.^>^.SV.NN\\\N\N ^^^^^S^-^ ::-:!-::-::-a' '■••'''■^■'■'•'■•■•■■■■■•^ [3 Post-storm ^ Pre-storm 20-30 30-40 40-50 -T" 50 —I 250 ... ,1 ,',\\M ■v:-:;-;--v-v-?':t T^:^^ ■!^-5WiWi«>^ S3 i^'S^^^V^ 150 200 250 0 Plankton volume {m!/1000m3) ~T 1 1 1 1 1 I I CO 200 300 400 500 600 700 Figure 8 Plankton volumes (miyiOOO m'') in MESSHAI tows. Average volumes from (A) deep tows and (B) shallow tows at inshore (solid I and offshore (shaded) stations; (C) volumes in 0-200 m stratum (average of pooled 40-m strata) from sequential deep tows at inshore station; (Dl volumes in 0- 40 m stratum (average of pooled 10-m strata) from sequential shallow tows at inshore station; average from (E) deep tows and (F) shallow tows at inshore station before storm (cross-hatching) and after storm (hatching). inshore station than at the offshore station (Fig. IIA; ANOVA of shallow strata, F^ 95=31.78, P=0.003). Within the upper 50 m at the inshore station, aver- age density was highest in the 10-20 m stratum, accounting for approximately AO^c of larvae. Density in the surface layer was comparatively low. In deep strata no significant difference in larval density was found between inshore and offshore stations (^1 ]7=0.53, P=0.47). At the inshore station, average larval densities were lower in poststorm samples than in prestorm samples (shallow strata: J'^j 54= 11.21, P=0.001; deep strata: Fj jo=10.05, P=0.06l). Overall, larval distribution was somewhat shallower after the storm (Fig. 11, E and F). Leuroglossus stilbius larvae In deep MESSHAI tows, larvae of L. stilbius occurred in relatively high densities in most strata down to 200 m (Fig. 12A). The lower limit of their distribution was not deter- mined by the MESSHAI samples, but another study 928 Fishery Bulletin 97(4), 1999 a A 0-40 1 40-80 1 ■ 80-120 120-160 1 160-200 1 1 1 1 1 1 B 0-0 16 0-10 10-20 20-30 30-40 40-50 100 200 300 400 500 600 700 800 ~! I I I 1 I I I I I 28 31 34 37 40 43 46 49 52 55 58 61 30 33 36 39 42 45 48 51 54 57 60 63 Tow number _ ?^^-»«'»>»W:«x<<^<«<^^ 0-0 16 0-10 10-20 20-30 30-40 40-50 >^woo^»cw}>c>ccic<>ooco<^CT ^s^..^^^^. ~\ 1 1 1 I 1 1 I 0 100 200 300 400 500 600 700 800 — r- 500 Eggs/ 100m3 Figure 9 Anchovy eggs (no./lOO m'M in MESSHAI tows. Average from lA) deep tows and (Bi shallow tows at inshore I solid i and offshore (shaded I stations; (C) densities in 0-40 m stratum from sequen- tial deep tows at inshore station; (Di densities in 0-40 m stratum (average of pooled 10-m strata! from sequential shallow tows at inshore station; average from (El deep tows and (F) shallow tows at inshore station before storm (cross-hatching) and after storm (hatching). (Moser and Smith, 1993) showed that they may oc- cur down to 300 m. Average densities were generally higher at the inshore station than at the offshore station. Larvae were absent from the 0-40 m stra- tum offshore and were nearly absent from this stra- tum in prestorm inshore samples (Fig. 12, A and E). In samples at the inshore station, average densities following the storm increased markedly at 0-80 m, decreased at 80-160 m, and increased slightly in the 160-200 m stratum (Fig. 12E). Larvae were not taken in Manta tows nor in the upper 40 m of shallow MESSHAI tows at the offshore station (Fig. 12B). At the inshore station the distribution was shifted up- ward in poststorm samples and average larval den- sities were relatively high in strata as shallow as 30-40 m (Fig. 12F). Genyonemus lineatus larvae White croaker larvae were absent from offshore samples and were abun- dant only in the upper 30 m at the inshore station (Fig. 13, A and B). Average densities were extremely low in prestorm samples but were relatively high in Moser and Pommeranz: Distribution of eggs and larvae of Engraulis mordax 929 Table 1 Taxa captured by Manta and MESSHAI nets on Cruise 8003-EB. Numbers of lai-vae are unadjusted counts; occurrences are listed in parentheses with onshore occurrences to the left of the si ash and offshore occurrences to the right. Only taxa identified to | genus or species are included. Num her of larvae and occurrences Shallow Deep Taxon Family Manta MESSHAI MESSHAI Total Engraulis mordax Engraulidae 3,066 (24/15) 67,640 (24/17) 19,696 (12/7) 90,402 99) Leuroglossus stilbius Bathylagidae 0 785 (22/2) 1,401 (12/7) 2,186 43) Genyonemus lineatus Sciaenidae 49 (7/0) 740 (19/0) 173 (8/0) 962 34) Stenobrachius leucopsarus Myctophidae 1 (0/1) 349 (24/9) 193 (12/7) 542 53) Sehastes spp. Sebastidae 53 (3/0) 264 (22/12) 134 (10/7) 441 54) Seriphus politus Sciaenidae 17 (6/0) 164 (12/0) 106 (5/0) 287 23) Peprilus siniillimus Stromateidae 6 (4/0) 180 (21/2) 35 (10/0) 221 37) Paralichthys californicus Paralichthyidae 9 (3/0) 82 (16/0) 23 (6/0) 114 25) Citharichthys spp. Paralichthyidae 0 63 (17/1) 14 (7/1) 77 26) Merluccius productus Merlucciidae 0 36 (1.5/0) 12 (5/4) 48 24) Atherinopsis californiensis Atherinidae 44 (9/0) 0 0 44 9) Pleuronichthys i-erticalis Pleuronectidae 1 (1/0) 25 (9/0) 8 (3/0) 34 13) Lyopsetta exitis Pleuronectidae 0 27 (16/0) 5 (4/0) 32 20) Bathylagus ochotensis Bathylagidae 0 21 (13/0) 10 (4/4) 31 21) Parophrys vetulus Pleuronectidae 0 19 (9/0) 7 (5/0) 26 14) Cataetyx rubnrostris Bythitidae 0 20 (12/0) 4 (3/0) 24 15/0) Argentina sialis Argentinidae 0 9 (8/0) 8 (6/0) 17 14) Protomyctophum crockeri Myctophidae 0 1 (1/0) 10 (4/4) 11 9) Argyropelecus spp. Sternoptychidae 0 0 9 (4/0) 9 4) Pleuronichthys spp. Pleuronectidae 0 6 (3/0) 3 (3/0) 9 6) Ophidian scrippsae Ophidiidae 0 1 (1/0) 6 (3/0) 7 4) Atractoscion nobilis Sciaenidae 0 6 (2/0) 0 6 2) Neoclinus stephensae Chaenopsidae 6 (2/0) 0 0 6 2) Hypsoblennius spp. Blenniidae 3 (2/0) 2 (2/0) 0 5 4) Pleuronichthys coenosus Pleuronectidae 4 (3/1) 1 (1/0) 0 5 5) Coryphopterus nicholsii Gobiidae 0 4 (3/0) 0 4 3) Scorpaenichthys marmoratus Cottidae 4 (2/0) 0 0 4 2) Tnphoturus mexicanus Myctophidae 0 1 (1/0) 3 (2/0) 4 3) Bathylagus wesethi Bathylagidae 0 3 (2/0) 0 3 2) Chauliodus macouni Chauliodontidae 0 0 3 (2/1) 3 3) Hypsopsetta guttulata Pleuronectidae 0 3 (3/0) 0 3 3) Lampanyctus spp. Myctophidae 0 1 (1/0) 2 (1/1) 3 3) Oxyjulis californica Labridae 0 3 (2/0) 0 3 2) Tarletonbeania crenulans Myctophidae 0 2 (2/0) 1 (1/0) 3 3) Zaniolepis frenata Hexagrammidae 0 2 (2/0) 1 (1/0) 3 3) Girella nigricans Kyphosidae 0 2 (2/0) 0 2 2) Pleuronichthys ritteri Pleuronectidae 0 2 (1/1) 0 2 2) Tactostoma macropus Melanostomiidae 0 0 2 (1/0) 2 1) Bathylagus millen Bathylagidae 0 1 (1/0) 0 1) Brosmophysis margmata Bythitidae 0 1 (1/0) 0 1) Chilara taylori Ophidiidae 0 1 (0/1) 0 1) Cololabis saira Scomberesocidae 1 (0/1) 0 0 1) Hypsoblennius genlilis Blenniidae 1 (1/0) 0 0 1) Hypsoblennius jenkmsi Blenniidae 1 (1/0) 0 0 1) Lampanyctus ritteri Myctophidae 0 1 (0/1) 0 1) Melamphaes sp. Melamphaidae 0 1 (1/0) 0 1) Scomber japonicus Scombridae 0 1 (1/0) 0 11 Trachipterus altivelis Trachipteridae 0 1 (l/Oi 0 1) Typhlogobius californiensis Gobiidae 1 (1/0) 0 0 1) 930 Fishery Bulletin 97(4), 1999 25000 -I 20000 15000 10000- 5000 0 1 jllUlM^ Vi'i'iTi-i n 1 1 1 1 1 2 3 4 5 6 7 8 9101112131415>15 2 3 4 5 6 7 8 9101112131415>15 2 3 4 5 6 7 8 9101112131415>15 360- 300- 250- F 200- 150- ■ 100- 1 50- 0- 1 ■ - 1,5 2.5 3.5 4.5 ' 5.5 ' Length (mm) Figure 10 Length frequencies of larvae of the 10 most abundant taxa taken in MESSHAI (A and D-Li. iB) oblique bongo, and (Cl Manta tows on Cruise 8003-EB. (A-C) Engraulis mordax; day (shaded! and night (solid) are shown for Manta tows; (Dl Leuroglossus stilhtus: (El Genyonemus lineatus; (F) Stenobrachius leucopsanis: (Gl Sebastes spp.; (H) Seriphiis politus; (II Pepnlus simillnniis; (Jl Paralichthys californicus; (Kl Cithanchthys spp.; (L) Merluccius productus. In A-C, bars represent 0.5-mm size classes; in D-L, bars represent the midpoints of 1.0-mm size classes. poststorm samples from the surface to 30 m (Fig. 13, E and F). Individual tows showed a sharp increase in lar- val density immediately following the storm, peaking 2 days after the storm and declining to prestorm levels 3 days after the storm (Fig. 13, C and D). Stenobrachius leucopsarus larvae Lai-vae of north- ern lampfish occurred throughout the water column to 200 m depth at the inshore station, with highest average densities in the 20-30 m and 30-40 m strata (Fig. 14, A and B ). Average densities were greater at the inshore station than at the offshore station. In deep MESSHAI tows taken at the offshore station, average density was greatest in the 40-80 m stra- tum; larvae were absent between 80 and 160 m. At the inshore station, maximum prestorm larval den- sities were in the 0-40 m and 40-80 m strata; in poststorm samples, densities were highest at 0-40 m ( Fig. 14E ). The center of distribution of larval densi- ties from shallow MESSHAI tows shifted upward from 30-40 m in prestorm samples to 20-30 m in poststorm samples (Fig. 14F). Individual deep Moser and Pommeranz: Distribution of eggs and larvae of Engraulis mordax 931 E o B 120-160 160-200 0-0 16 0-10 10-20 20-30 30-40 40-50 400 600 800 1000 1200 1400 0 500 Larvae / 100m3 — I — 1000 29 32 35 38 41 44 47 50 53 56 59 62 ^ss^^^^^^^^^a M I I I I I I I M 28 31 34 37 40 43 46 49 52 55 58 61 30 33 36 39 42 4S 48 51 54 57 60 63 Tow number _ £ 40-80 i 2 80-120 U) 120-160 c<:<'yMC<<>&ox<>oiOOi6Mi6fMi(fy>6iii66/i6ia 0 500 1000 1500 2000 2500 0 500 1000 1500 2000 2500 3000 3500 Larvae/ 100m3 Figure 11 Anchovy larvae (no./lOO m3) in MESSHAI tows. Average from (A) deep tows and (B) shallow tows at inshore (solid) and offshore (shaded) stations; (C) densities in 0-40 m stratum from sequential deep tows at inshore station; (Dl densities in 0-40 m stratum (average of pooled 10-m strata) from sequential shallow tows at inshore station; average from (E) deep and (F) shallow tows at inshore station before storm (cross-hatching) and after storm (hatching). MESSHAI tows in the upper 200 m showed a trend of increasing density just before the storm and a sharp decrease in poststorm tows (Fig. 14C). In shal- low MESSHAI tows, densities peaked in two tows taken one day before the storm but were not appreciably dif- ferent in pre- and poststorm samples (Fig. 14D). Sebastes spp. larvae Rockfish larvae were concen- trated in the upper 80 m of the water column, with low average densities between 80 and 200 m (Fig. 15A). In deep MESSHAI tows the highest average density was in the 40-80 m stratum at the offshore station but in the 0-40 m stratum at the inshore sta- tion (Fig. 15A). In shallow MESSHAI tows at the inshore station, average density was highest in the 20-30 m stratum, relatively low in the 0-10 m and 10-20 m strata, and nearly as high at the surface as in the 20-30 m stratum. Offshore, shallow MESSHAI tows showed extremely low average densities in the upper 30 m of the water column; larvae were not taken in surface net tows. At the inshore station, average larval densities were higher in samples 932 Fishery Bulletin 97(4), 1999 a 0^0 A 40-80 1 - 80-120 120-160 1 160-200 1 C 1 1 1 1 1 50 100 150 200 250 B 0-0 16 0-10 10-20 I 20-30 ■ 30-40 40-50 — 1 1 1 1 T '1 0 50 100 150 200 250 300 350 Larvae/ 1000m3) 250 T 29 32 35 38 41 44 47 50 53 56 59 62 28 31 34 37 40 43 46 49 52 55 58 61 30 33 36 39 42 45 48 51 54 57 60 63 Tow number .§. 40-80 E 2 80-120 0-016 VVWWWM i«^«<'«<<<<«»>X«-X«M wowwk^«k»i^ LVjiitjMiji'jiiiiVi'i.ii^ri I iVi^j < 1^ || — ■ — " ^3^ 10-20 20-30 30-40 40-50 I ■ ' ■ ■ I ;, ' 1 250 300 0 100 Larvae / 1000m3 -r Figure T2 Leuroglossus stilbius larvae (no./lOOO m-'i in MESSHAI tows. Average from lAi deep and (B) shallow tows at inshore (solid) and offshore (shaded) stations; (C) densities in 0-200 m stratum (average of pooled 40-m strata) from sequential deep tows at inshore station; (D) densities in 0- 40 m stratum (average of pooled 10-m strata) from sequential shallow tows at inshore station; average from (E) deep and iF) shallow tows at inshore station before storm (cross-hatching) and after storm (hatching). taken before the storm than those taken afterwards; however, the disparity between average densities be- fore and after the storm was due largely to several prestorm tows with unusually high larval counts (Fig. 15, C-F). Seriphus politus larvae Queenfish larvae occurred only at the inshore station, primarily in the upper 10 m of the water column (Fig. 16, A and B). At the inshore station, larvae were virtually absent from tows taken before the storm; larval density increased abruptly immediately following the storm, then sub- sided to prestorm level within about two days after the storm (Fig. 16, C-F). Peprilus simillimus larvae Larval Pacific pompano occurred primarily in the upper 40 m, with highest average densities in the upper 20 m of the water col- umn (Fig. 17, A and B). Except for two tows at the 30-40 m stratum, larvae did not occur in offshore Moser and Pommeranz: Distribution of eggs and larvae of Engraulis mordax 933 29 32 35 38 41 44 47 50 53 56 59 62 28 31 34 37 40 43 46 49 52 55 58 61 30 33 36 39 42 45 48 51 54 57 60 63 Tow number _ t, ' 1 ^ 40-80 E ra 80-120 9- 120-160 a SX3 —I — 100 200 250 0 100 200 300 400 500 Larvae/ 1000m3 Figure 13 Genyonemus lineatus larvae (no./lOOO m^l in MESSHAI tows. Average from (A) deep and (Bi shallow tows at inshore (solid) and offshore (shaded) stations; (C) densities in 0-40 m stratum from sequential deep tows at inshore station; (D) densities m 0-40 m stratum (average of pooled 10-m strata) from sequential shallow tows at inshore station; average from (E) deep and (F) shallow tows at inshore station before storm (cross-hatching) and after storm (hatching). samples. At the inshore station, lai-val densities were comparatively low in prestorm samples but increased markedly after the storm (Fig. 17, C-F). Paralichthys californicus larvae California halibut larvae occurred only at the inshore station and were restricted to the upper 30 m of the water column ( Fig. 18, A and B). Average larval density was highest in the 0-10 m stratum, where it was four times higher than in the 20-30 m stratum or at the surface (Fig. 18B). Larvae appeared in several of the shallow MESSHAI tows prior to the storm at densities as high as approximately 20 per 1000 m^; after the storm, densities increased steadily to more than 60 per 1000 m'^, then decreased to prestorm levels within 2 days following the storm (Fig. 18D). Average poststorm larval density for the 10-m stratum was more than 7 times higher than prestorm density, and overall vertical distribution was slightly shallower in poststorm samples (Fig. 18F). Larvae occurred in Manta net samples only in poststorm tows. 934 Fishery Bulletin 97(4), 1999 E o B 70- 60- 50- 40- 30- 20- 10- 0- 0-0 16 0-10 10-20 20-30 30-40 40-50 ~l 1 1 1 1 1 1 1— I 10 20 30 40 50 60 70 80 90 Larvae/ 1000m3 250- 200- 100- D T — I— 1—1 — I I I I — I — I — r 29 32 35 38 41 44 47 50 53 56 59 62 I I I I I I I I 28 31 34 37 40 43 46 49 52 55 58 61 30 33 36 39 42 45 48 51 54 57 60 63 Tow number _ :k<<>?ixx>»x<>x>c>c»x<<^^ Q- 120-160 •, E^ •120 ^^^^ ^^^^^22 ^^^■:-l.;.:-:.;-:.:-:-n 0-0 16 0-10 10-20 20-30 30-40 40-50 VVVVVM ■:■;■^:■;^;^^^:■;■;;:■^:^:■^^^■M XX3 yJ'iiM.Wiii'iMMiiiMiJ'rt^^ xiKi(AyA 30 40 50 60 -I 1 1 T ^ \ r^n 20 40 60 80 100 120 140 160 Larvae/ 1000m3 Figure 14 Stenobrachius leucopsarus larvae (no./lOOO ni ') in MESSHAI tows. Average from (A) deep and iBi shallow tows at inshore Isolid) and offshore (shadedi stations; (C) densities in 0-200 m stra- tum (average of pooled 40-m strata) from sequential deep tows at inshore station; (Dl densities in 0-40 m stratum (average of pooled 10-m strata) from sequential shallow tows at inshore sta- tion; average from (E) deep and (F) shallow tows at inshore station before storm (cross-hatching) and after storm (hatchingl. Citharichthys spp. larvae Sanddab larvae were taken primarily in the upper 50 m of the water col- umn at the inshore station ( Fig. 19, A and B ). In shal- low MESSHAI tows, peak average density was in the upper 10 m. Poststorm samples contributed most of the larvae (Fig. 19, C-F). A few larvae appeared spo- radically in tows taken before the storm; following the storm there was a steady increase in larval den- sity and then a sharp decline by the second day after the storm (Fig. 19, C and D). The distribution was shifted upward by approximately 20 m in poststorm samples (Fig. 19F). Merluccius productus larvae Hake larvae were taken in deep MESSHAI tows in the 40-80 and 80-120 m strata; average densities were similar in inshore and offshore samples (Fig. 20A). Larvae occurred in shal- low MESSHAI tows only at the inshore station and were taken as shallow as 10-20 m ( Fig. 20B ). After the storm only one deep MESSHAI tow was positive, but for shal- Moser and Pommeranz: Distribution of eggs and larvae of Engraulis mordax 935 E o o o 29 32 35 38 41 44 47 50 53 56 59 62 ^^^fe^^^^W'W^K^^-'X^X^ ^ 40-80 E 2 80-120 ^»^W«^>6«^«>«>W^ 28 31 34 37 40 43 46 49 52 55 56 61 30 33 36 39 42 45 48 51 54 57 60 63 Tow number _ 016 h^>iM^Mi^,i^>i/XM66666iMi6(^^^ — t 60 ^^>WiWi6iiii6Mi6(^ 1 1 \ r— I r— I 1— I 10 20 30 40 50 60 70 80 90 Lawae/ 1000m3 Figure 15 Sebastes spp. larvae (no./lOOO m3) in MESSHAI tows. Average of (A) deep and (B) shallow tows at inshore (solid) and offshore (shaded) stations; (C) densities in 0-200 m stratum (average of pooled 40-m strata) from sequential deep tows at inshore station; (D) densities in 0-40 m stra- tum (average of pooled 10-m strata) from sequential shallow tows at inshore station; average from (E) deep and (F) shallow tows at inshore station before storm (cross-hatchmg) and after storm (hatching). low MESSHAI tows, the frequency of positive tows in- creased, along with larval densities (Fig. 20, C and D). The distribution of larvae became somewhat shallower in shallow MESSHAI tows after the storm (Fig. 20F). Discussion Northern anchovy larvae dominated samples from both study sites. The midwater species Leuroglossus stilbius and Stenobrachius leiicopsarus and the rock- fish genus Sebastes were relatively abundant at the two sites. These three taxa, along with Merluccius productus (ranked 10th in abundance) and Bathy- lagus ochotensis (ranked 13th), constitute a distinct larval recurrent group in the California Current re- gion ( Moser and Smith, 1993 ). These three taxa were also closely linked in other analyses of larval fish assemblages in the study region (Gruber et al., 1982; McGowen, 1993). Sebastes spp. and Merluccius 936 Fishery Bulletin 97(4), 1999 i 60-120 B T 1 1 1 1 1 1— f 20 40 60 80 100 120 140 160 D 0 'Tff f f!T!fTlf 28 31 J4 37 40 43 46 49 52 5S 68 61 30 33 36 39 42 45 48 51 54 57 60 63 Tow number _ £ 40-80 E 3 2 80-120 ^ ^ \\\\\V\\\\VVV\\^\V\^-^V1 0-0 16 0-10 10-20 20-30 30^0 40-50 IS3 — I 1 1 \ 1 1 1 I 0 20 40 60 80 100 120 140 160 \\\\l 0 50 100 150 200 250 300 Larvae/ 1000m3 Figure 16 Senphiis politus larvae (no./lOOO m') in MESSHAl tows. Average from (A) deep and iBl shallow tows at inshore (solid) and offshore (shaded) stations; (C) densities in 0-40 m stratum from sequential deep tows at inshore station; (D) densities in 0-40 m stratum (average of pooled 10-ni strata! from sequential shallow tows at inshore station; average from (E) deep and (F) shallow tows at inshore station before storm (cross-hatching) and after storm (hatching). productus are demersal and their larvae are found throughout the SCB at this season (Moser et al., 1993 ). Other studies (Gruber et al., 1982; McGowen, 1993 ) have shown that larvae of Leuroglossus stilbius and Stenobrachius leucopsarus are relatively abun- dant nearshore and, according to Barnett et al. (1984), S. leucopsarus mayextend shoreward to midshelf. This can be explained, in part, by the nar- rowness of the shelf in this region and the proximity to deep-water habitat. These species are unusual in comparison with other bathylagids and myctophids in this region in having relatively high larval abun- dances near the coast (Moser et al., 1993). Most lar- vae of other taxa collected in this study were taken at the nearshore station and represent demersal shelf species. The most abundant of these, Genyonemus lineatus, spawns during winter and early spring over shelf waters throughout the SCB (Gruber et al., 1982 Schlotterbeck and Connally, 1982; Watson, 1982 Barnett et al., 1984; Brewer and Kleppel, 1986 Lavenberg et al., 1986; Walker et al., 1987; McGowen, 1993; Moser et al., 1993). Seriphus politus. the next Moser and Pommeranz: Distribution of eggs and larvae of Engraulis mordax 937 -r- 25 B 0-0 16 0-10 - 10-20 20-30 MM 30^10 i 40-50 1 1 30 0 10 20 30 40 50 60 70 80 Larvae/ 1000m3 I I I I I I I I I I 28 31 34 37 40 43 46 49 52 55 58 61 30 33 36 39 42 45 48 51 54 57 60 63 Tow number ^ 40-80 E 2 80-120 ,^.S\^.S^,^.\^^.\..^.\^^SSSSN^ -1— 10 1^ 25 0-0 16 F 0-10 > \ V 'x \ \ \ \ \ '\ V N V V V \ \ ■- '^1 ^ ^ 10-20 \\\\\\\\\\\\\\\\\'\'^:\\\\ yiwk ,\V\M 30^0 40-50 1 1 1 1 1 1 1 30 0 20 40 60 80 100 120 140 Larvae/ 1000m3 Figure 17 Peprilus simUUmus larvae (no./lOOO niM in MESSHAI tows. Average from ( Al deep and (B) shal- low tows at inshore (solid) and offshore (shaded! stations; (C) densities in 0-40 m stratum from sequential deep tows at inshore station; (D) densities in 0-40 m stratum (average of pooled 10-m strata) from sequential shallow tows at inshore station; average from lE) deep and (F) shallow tows at inshore station before storm (cross-hatching) and after storm (hatching). most abundant shelf species in the samples, typically spawns from late spring to summer; however, spawn- ing began unusually early in 1980 since larvae were abundant as early as March off San Onofre (Walker et al., 1987). Peprilus simillimus, Par-alichthys cali- fornicus, and Cithanchthys spp. have broad spawn- ing seasons with high larval abundance in late win- ter to early spring. This was particularly evident in .1980 (Walker et al., 1987; Moser et al., 1993). The vertical distribution of larval fishes is closely related to the temperature profile of the water column. Ahlstrom ( 1959) described two general categories: those taxa whose larvae occur almost entirely within the upper mixed layer and in the upper part of the ther- mocline; and those that occur within or below the ther- mocline. Larvae of shorefishes, including most clupeoids, typically fall within the first category. Gen- erally, slope and offshore taxa produce deeper-living larvae, although many oceanic taxa also have shallow- living larvae. Northern anchovy had similar vertical profiles at the slope and offshore stations although den- sities were much higher at the slope station, where the 938 Fishery Bulletin 97(4), 1999 29 32 35 38 41 44 47 50 53 56 59 52 E 0-40 40-60 \\\\\\\\\\VW\\\\\\\V\1 ««««*t«4«*3 E, e 80-120 Q. 120 160 Q 160-200 28 31 34 37 40 43 46 49 52 55 58 61 30 33 36 39 42 45 48 51 54 57 60 63 Tow number 00 16 0-10 10-20 20-30 30^0 40-50 [•■-• • • •! 55SSSSSSSSXX3SSS3SSS3 5S?SXSSSSXS3 T- 10 I 40 — r- 50 — r- 70 Larvae / 1000m3 Figure 18 Paralichthys californicus larvae (no./lOOO m*) in MESSHAI tows. Average from ( Al deep and (Bi shallow tows at inshore (solid) and offshore i shaded i stations; iC) densities in 0-40 m stratum from sequential deep tows at inshore station; (D) densities in 0-40 m stratum (average of pooled 10-m strata) from sequential shallow tows at inshore station; average from (E) deep and (Fi shallow tows at inshore station before storm (cross-hatching) and after storm (hatching). density peak was more pronounced and was about 10 m shallower than at the offshore station. Probably this peak was related to the relatively shallow mixed layer at the slope station compared with the offshore sta- tion. Eggs of the anchovy had a shallower distribution than the larvae, with an apparent density peak in the neuston. The effect of the different mixed layer depths at the two stations is evident in the slightly deeper lower distribution boundary of eggs at the offshore station. Like the anchovy, the nearshore shelf species (white croaker, queenfish, California halibut, California pom- pano, and some sanddab species) had shallow larval distributions essentially limited to the upper 30 m of the water column of the inshore station, with peak den- sities in either the 0-10 m or 10-20 m strata. Such shallow distributions appear to be typical of nearshore species although previous studies have lacked the fine- scale sampling capability needed to show this. The many Sebastes species occupy a wide range of habitats on the shelf and upper slope and produce larvae that are found within and below the thermocline and that generally avoid the upper mixed layer, except for the Moser and Pommeranz: Distribution of eggs and larvae of Engraulis mordax 939 A " 40-80 E i= 80-120 O 120-160 o n ID 0-0-16 0-10 B 10-20 20-30 ^^^1 ^H 30^0 40-50 A 0 1 5 1 10 1 15 1 20 25 Larvae/ 1000m3 fiO-i D 50- Storm — > l\ 30- J\ 10- 0- w^ Vv- 29 32 35 38 41 44 47 50 53 56 59 62 0-40 ^ -E 40-80 E 2 80-120 p. 120-160 (1) D \'>,^^^^^^^^v^:A^ 28 31 34 37 40 43 46 49 52 55 58 61 30 33 36 39 42 45 48 51 54 57 60 63 Tow number „ 0-0 16 0-10 }■-•-■ • • -• •■•-• •■' •! 10-20 § 20-30 30^0 40-50 ■ ' ■ ■ ' ■ 1 ^ I I I \ 1 1 1 1 1— I 0 12 3 4 5 6 7 8 9 0 5 10 15 20 25 30 35 40 45 50 Larvae/ 1000m3 Figure 19 Citharichthys spp. larvae (no./lOOO mJ) in MESSHAI tows. Average from (A) deep and (B) shal- low tows at inshore (solid) and offshore (shaded) stations; (C) densities in 0-80 m stratum (aver- age of pooled 40-m strata) from sequential deep tows at inshore station; (D) densities in 0-40 m strata (average of pooled 10-m strata) from sequential shallow tows at inshore station; average from (El deep and (F) shallow tows at inshore station before storm (cross-hatching) and after storm (hatching). neustonic zone (Ahlstrom, 1959; Moser and Boehlert, 1991). This distribution is clearly shown by comparing the stratum of highest density at the inshore and off- shore stations. Likewise, Pacific hake are broadly dis- tributed over the shelf and slope and have deep-living larvae (Ahlstrom, 1959; Mullin and Cass-Calay, 1997). In our study, hake larvae were absent from samples above 50 m depth at the offshore station. A comparison of larval anchovy length distribu- tions from MESSHAI, bongo, and Manta net tows shows that the MESSHAI is less effective in catch- ing larvae over 3 mm than the other two (Fig. 10, A-C). The mouth opening of the MESSHAI net is relatively much smaller (25x25 cm) than that of the bongo net ( 7 1 cm diameter). Why the Manta net cap- tures larger larvae is conjectural. It may be related to the absence of an upward escape plane for larvae in the path of the net or it may reflect the size fre- quency of larvae in the surface layer. The bimodal length distribution for anchovy captured in Manta 940 Fishery Bulletin 97(4), 1999 Q. Q A 0-40 - 40-80 ^^ 80-120 120-160 150-200 1 1 1 1 1 1 1 I — I B 0-016 0-10 10-20 20-30 30-40 40-50 0 0 5 10 15 20 25 30 35 40 46 ~1 1 1 1 1 \ 1 1 012345678 4.5- 4.0- 35- 30- 25- 20- 15- 1 0- 05- 0- Larvae/1000m3 25- 20- 29 32 35 38 41 44 47 50 53 56 59 62 / I T T t ° IfTf lYitTT «<<<<<<>>>x<<>>:<<>>>>>>>>>:c>>:<>>>>>>>>>>> 28 31 34 37 40 43 46 49 52 55 58 61 30 33 36 39 42 45 48 51 54 57 60 63 Tow number _ 0-0 16 2 8o-'2c ^«^>^j<^^i«>^^^«■;;?^ 0-10 10-20 20-30 VV\\VVV\V1 .Vio^■^^^^^^\^\\\^^^^:sSl 30-40 .\>aA.\^«\).^A^vi.^^^^^^^M ■50 '^^,^!^6(>i!ii^iiy>6Mi(M>!>^ 7 0 2 Larvae/ 1000m3 -T- 10 —I 12 Figure 20 Merluccius productus larvae (no./lOOO m') in MESSHAI tows. Average from (A) deep and (B) shallow tows at inshore (solid) and offshore (shaded) stations; (C) densities in 0-200 m stratum (average of pooled 40-m strata) from sequential deep tows at inshore station; (D) densities in 0- 40 m stratum (average of pooled 10-m strata) from sequential shallow tows at inshore station; average from (E) deep and (F) shallow tows at inshore station before storm (cross-hatching) and after storm (hatching). tows results from relatively large numbers of 6.0- 10.0 mm larvae caught at the surface during the day (Fig. IOC). This was a consistent feature of Manta catches throughout the cruise, suggesting that an- chovy larvae may have been migrating to the sur- face layer to feed during the day. The storm that interrupted sampling for two days at the inshore station significantly affected the rela- tive abundance and distribution of fish larvae and their physical and biotic environment. Average den- sities of anchovy larvae declined markedly in most strata following the storm, possibly as a result of mortality associated with starvation. This argument is contradicted by the results of Mullin et al. (1985) who showed that important larval fish prey, such as copepod nauplii increased in concentration after the storm, were no less stratified than before the storm, and had a 15-m upward shift in peak abundance that mirrored the shoaling of anchovy larval distribution in our study. Moreover, the chlorophyll maximum was Moser and Pommeranz Distribution of eggs and larvae of Engraulis mordax 941 below the peak depth zone of anchovy larval abun- dance before the storm, whereas it coincided with the depth zone of highest larval abundance after- wards. Pre- and poststorm peak concentrations of important fish larva prey coincided with pre- and poststorm peaks in the chlorophyll maximum (MuUin et al., 1985). If starvation was the reason for the de- cline in anchovy larvae, then it is likely that the cause was disruption of micropatches of food, as suggested by Mullin et al. (1985). In this case, Lasker's (1981) stable ocean hypothesis was operative at the centi- meter scale, and reduced survival was a result of dis- ruption of the fine-scale geometry of food patches (Vlymen, 1977; Owen, 1989).Another explanation for the poststorm decline in anchovy larval abundance would be advection of larvae away from the study site (Mullin etal., 1985). The strong northwest winds resulted in offshore movement of surface water (the upper 10 m) by Ekman transport at a rate of about 10 cm per second; thus, virtually all the surface wa- ter at this region of the shelf was moved offshore during the storm and was replaced by deeper water (WinantM. This could explain the sharp decline in anchovy larval densities at the slope station after the storm, providing, of course, that densities of an- chovy larvae were lower over the shelf than at the station before the storm. The latter would not be expected since Barnett et al. ( 1984 ) showed high con- centrations of anchovy larvae over the shelf at a simi- lar habitat south of our inshore station. The storm had an opposite effect on densities of shorefish larvae (Genyonemus lineatus, Seriphus politus, Pepriliis simillimiis, Paralichthys califor- nicus, Citharichthys spp.), which occurred in ex- tremely low densities or which were absent in prestorm samples. Larval densities of these species began to increase immediately after the storm, peaked within 1-2 days, and then declined abruptly to prestorm levels. The sudden appearance of these larvae at the nearshore station after the storm was a result of storm-induced advection from shallow regions of the shelf The narrow shelf in this region of the coast would make nearshore fish lai"vae par- ticularly vulnerable to transport off the shelf during storms with northwest winds. Such advection may be important in the transport of nearshore fish lar- vae seaward where they become available to slope or eddy circulation, thus providing the opportunity for dispersion to other regions. The subsequent rapid decline in larval densities could have been caused by mortality associated with disturbance of micro- ti C. D.Winant. 1998. Center for Coastal Studies, Scripps In- stitution of Oceanography, La JoUa, CA 92093. Personal commun. patches of food, by predation, or by active or passive movement of larvae away from the sampling site. Starvation of these inshore larvae over slope waters may be related to the difference in composition and concentration of prey organisms in nearshore and offshore waters (Watson and Davis, 1989). Some por- tion of the larvae that appeared at the inshore site after the storm may have been transported back to the nearshore region. This return transport may have been enhanced by upward movement of these larvae to the surface where they could be carried by shore- ward currents created by internal wave cells (Shanks, 1983, 1986). Among the species with deeper-living larvae, those of Sebastes spp. declined markedly in all strata after the storm, whereas those of Stenobrachiiis leucop- sarus showed a decline in deeper strata but increased in strata shallower than 30 m. This probably was caused by upward turbulent advection of larvae from deeper strata. Likewise, larval density of Leuro- glossus stilbius increased in strata shallower than 50 m following the storm, with sporadic high counts in individual tows, suggesting pulses of upwardly advected larvae. Acknowledgments We are indebted to the members of the scientific party of Cruise 8003-EB (Jack Brown, John Butler, Carol Kimbrell, Barbara Sumida-McCall, Elaine Sandknop Acuha, and Elizabeth Stevens) for their outstanding work in all phases of cruise operations. The efforts of Captain L. E. Davis and the crew members of the RV Ellen B. Sc/'ipps were crucial to the completion of the cruise objectives. The late J. T Thrailkill su- pervised sorting of the plankton and processing of the fish eggs and larvae. Barbara Sumida-McCall, Elaine Sandknop Acuiia, and Elizabeth Stevens staged the anchovy eggs and identified and measured fish larvae. We are indebted to Carol Kimbrell for directing the chlorophyll sampling at sea and to Rob- ert Owen for his help in working up the chlorophyll data from these samples. Richard Charter assisted us in all aspects of data processing associated with this study. William Watson checked identifications of some nearshore fish families and offered sugges- tions that improved the manuscript. We thank Michael Mullin for reading the manuscript and for his comments and suggestions. Discussions with Clinton Winant were helpful in understanding nearshore circulation and transport. We are grate- ful to Nancy Lo for her review of the manuscript and for statistical procedures. The comments and sug- gestions of three anonymous reviewers greatly im- 942 Fishery Bulletin 97(4), 1999 proved the manuscript. This paper is dedicated to the memory of our friend and mentor, Reuben Lasker, who wholeheartedly supported this study and pro- vided encouragement and guidance to all of us. Literature cited Ahlstrom, E. H. 1959. Vertical distribution of pelagic fish eggs and larvae offCalifornia and Baja California. Fish. Bull. 60:107-146. Allen, M. J., and K. T. Herbinson. 1991. Beam-trawl survey of l:)ay and nearshore fishes of the soft-bottom habitat of Southern California in 1989. Calif Coop. Oceanic Fish. Invest. Rep. 32:112-127. Arnold, E. L. 1959. The Gulf V Plankton Sampler. U.S. Fish. Wildl. Serv. Circ. 62:111-113. Barnett, M. A., A. E. Jahn, P. D. Sertic, and W. Watson. 1984. Distribution of ichthyoplankton off San Onofre, Cali- fornia, and methods for sampling very shallow coastal waters. Fish. Bull. 82:97-111. Brewer, G. D., and G. S. Kleppel. 1986. Diel vertical distribution offish larvae and their prey in nearshore waters of southern California. Mar. Ecol. Prog. Ser. 27:217-226. Brown, D. M., and L. Cheng. 1981. New net for sampling the ocean surface. Mar. Ecol. Prog. Ser. 5:225-227. Checkley, D. M., Jr., P. B. Ortner, L. R. Settle, and S. R. Cummings. 1997. A continuous underway fish egg sampler. Fish. Oceanogr. 6:58-73. Cross, J. N. 1987. Demersal fishes of the upper continental slope off Southern California. Calif Coop. Oceanic Fish. Invest. Rep. 28:15.5-167. Cross, J. N., and L. G. Allen. 1993. Fishes. /;; M. D. Dailey, D. J. Reish, and J. W. Ander- son leds. 1. Ecology of the Southern California Bight: a syn- thesis and interpretation, p. 459-540. Univ. Calif Press, Berkeley. CA. Ebeling, A. W., R. M. Ibara, R. J. Lavenberg, and F. J. Rohlf. 1970. Ecological groups of deep-sea animals off Southern California. Bull. Los Angeles Cty. Mus. Nat, Hist. 6, 43 p. Eppley, R. W. (ed.l. 1986. Plankton dynamics of the Southern California Bight. In M. J. Bowman, R. T. Barber, and C. N. K. .Moocrs (eds.). Lecture notes on coastal and estuarine studies 15. Springer- Verlag, New York, NY. Feder, H. M., C. H. Turner, and C. Limbaugh. 1974. Observations on fishes associated with kelp beds in Southern California. Calif Dcp. Fish Game Fish Bull. 160, 144 p, Gruber, D., E. H. Ahlstrom, and M. M. Mullin. 1982. Distribution of ichthyophuiklon in the Southern Cali- fornia Bight. Calif, (.'oop. Oceanic Fish. Invest. Rep. 23:172-179. Hobson, E. S., and J. R. Chess. 1976. 'I'rophic interactions among fishes and zooplankters near shore at Santa Catalina Island, California. Fish. Bull. 74:.567-598. Hunter, J. R., and N. C. H. Lo. 1993. Ichthyoplankton methods for estimating fish biomass, introduction and terminology. Bull. Mar. Sci. 53:723-727. Kramer, D., M. J. Kalin, E. G. Stevens, J. R. Thrailkill, and J. R. Zweifel. 1972. Collecting and processing data on fish eggs and lar- vae in the California Current region. U.S. Dap. Commer., NOAATech. Rep. NMFS Circ. 370, 38 p. Lasker, R. 1978. The relation between oceanographic conditions and larval anchovy food in the California Current: identifica- tion of factors contributing to recruitment failures. Rapp. P.-V. Reun. Cons. Int. Explor. Mer 173:212-230. 1981. Factors contributing to variable recruitment of the northern anchovy iEngraulis ntordax) in the California Current: contrasting years 1975 through 1978. Rapp. P.- V. Reun. Cons. Int. Explor. Mer 178:375-388. Lasker, R. (ed.) 1985. An egg production method for estimating spawning biomass of pelagic fish: application to the northern anchovy iEitfiraiilis mordax). U.S. Dep. Commer., NOAA Tech. Rep. NMFS 36, 99 p. Lasker, R., J. Palaez, and R. M. Laurs. 1981. The use of satellite infrared imagery for describing ocean processes in relation to spawning of the northern anchovy (Engraulifi mordax). Remote Sens. Environ. 11:439-453. Lavenberg, R. J., G. E. McGowen, A. E. Jahn, J. H. Petersen, and T. C. Sciarrotta. 1986. Abundance of southern California nearshore ichthyo- plankton. Calif Coop. Oceanic Fish. Invest. Rep. 27:53-64. Love, M. S., J. S. Stephens Jr., P. A. Morris, M. M. Singer, M. Sandhu, and T. C Sciarotta. 1986. Inshore soft substrata fishes in the Southern Cali- fornia Bight: an overview. Calif Coop. Oceanic Fish. In- vest. Rep. 27:84-106. Lynn, R. J., K. A. Bliss, and L. E. Eber. 1982. Vertical and horizontal distributions of seasonal mean temperature, salinity, sigma-t, stability, dynamic height, oxygen, and oxygen saturation in the California Current, 1950-1978. CalCOFI Atlas 30, 513 p. McGowen, G. E. 1993. Coastal ichthyoplankton assemblages, with empha- sis on the Southern California Bight. Bull. Mar. Sci. 53:692-722. Mearns, A. J. 1979. Abundance, composition, and recruitment of nearshore fish assemblages on the Southern California mainland shelf Calif Coop. Oceanic Fish. Invest. Rep. 20:111-119. Moser, H. G., and E. H. Ahlstrom. 1985. Staging anchovy eggs. //) R. Lasker (ed.l. An egg production method for estimating spawning biomass of pelagic fish: application to the northern anchovy lEngmiilis mordax), p. 37-41. U.S. Dcp. Commer, NOAATech. Rep. NMFS 36. Moser, H. G., and G. W. Boehlert. 1991. Ecology of pelagic larvae and juveniles of the genus Schastcs. Env Biol. Fish. 30:203-224. Moser, H. G., R. L. Charter, P. E. Smith, D. A. Ambrose, S. R. Charter, C. A. Myer, E. M. Sandknop, and W. Watson. 1993. Dislnbulional atlas of fish larvae and eggs in the California Current region: taxa with 1000 or more total larvae, 1951 through 1984. CalCOFI Atlas 31, 233 p. Moser, H. G., and P. E. Smith. 199.3. Larval lish assemblages of the California Current Moser and Pommeranz: Distribution of eggs and larvae of Engraulis mordax 943 region and their horizontal and vertical distributions across a front. Bull. Mar. Sci. 53:645-691. Mullin, M. M., E. R. Brooks, F. M. H. Reid, J. Napp, and E. F. Stewart. 1985. Vertical structure of nearshore plankton off South- ern California: a storm and a larval fish food web. Fish. Bull. 83:151-170. Mullin, M. M., and S. L. Cass-Calay. 1997. Vertical distributions of zooplankton and larvae of the Pacific hake (whiting), Merlucciiis productus, in the California Current system. Calif. Coop. Oceanic Fish. Invest. Rep. 38:127-136. Nellen, W., and G. Hempel. 1969. Versuche zur Fangigkeit des "Hai" und des modifizier- ten Gulf-V-Plankton-Samplers "Nachthai". Ber. Dt. Wiss. Komm. Meeresforsch. 20:141-154. Owen, R. W. 1989. Microscale and finescale variations of small plank- ton in coastal and pelagic environments. J. Mar. Res. 47:197-240. Pommeranz, T., and H. G. Moser. 1987. Data report on the vertical distribution of the eggs and larvae of northern anchovy, Engraulis mordax. at two stations in the Southern California Bight, March-April, 1980. U.S. Dep. Commer., NOAA Tech. Memo. NMFS SWFC-75, 140 p. Schlotterbeck, R. E., and D. W. Connally. 1982. Vertical stratification of three nearshore southern California larval fishes (Engraulis mordax. Genyonemus lineatus. and Senphus politus). Fish. Bull. 80:895-902, Shanks, A. L. 1983. Surface slicks associated with tidally forced internal waves may transport pelagic larvae of benthic invertebrates and fishes shoreward. Mar. Ecol. Prog. Ser. 13:311-315. 1986. Vertical migration and cross-shelf dispersal of larval Cancer spp. and Randallia ornata (Crustacea: Brachyura) ofl'the coast of southern California. Mar. Biol. 92:189-199, Shepard, F., and K. O. Emery. 1941. Submarine topography off the California coast: Can- yons and tectonic interpretations. Geol. Soc. Am. Spec. Pap. 31, 171 p. Vlymen, W. J. 1977. A mathematical model of the relationship between larval anchovy (Engraulis mordax) growth, prey distribu- tion, and larval behavior. Env Biol. Fish. 2:211-233. Walker, H. J., W. Watson, and A. M. Bamett. 1987. Seasonal occurrence of larval fishes in the nearshore Southern California Bight off San Onofre, California, Estuar. Coast. Shelf Sci. 25:91-109. Watson, W. 1982. Development of eggs and larvae of the white croaker, Genyonemus lineatus Ayres (Pisces: Sciaenidae). off the Southern California coast. Fish. Bull. 80:403-417. Watson, W., and R. L. Davis. 1989. Larval fish diets in shallow coastal waters off San Onofre, California. Fish. Bull. 87:569-591. 944 Abstract.— Growth parameter esti- mates were calculated for the tiger shark (Galeocerdo cuvier) by using tag and recapture data. Results were com- pared to published estimates based on bands in vertebrae. The von Bertalanffy parameters (sexes combined) based on tag and recapture growth data were as follows; L„ = 337 cm fork length, k - 0.178, and /o= -1.12. Monthly length- frequcmcy data for six year classes from birth to two years old for tiger sharks were used to verify the tag-recapture growth curve for this age range. The predicted age at maturity is 7 years for both sexes. Data from an ongoing in situ study with oxytetracycline were used in conjunction with length data to de- termine the effect of tagging and oxytet- racycline injection on growth. The data suggest that tagging alone or tagging combined with oxytetracycline injection has little or no effect on the growth rate of tiger sharks up to two years of age. Growth of the tiger shark, Galeocerdo cuvier, in the western North Atlantic based on tag returns and length frequencies; and a note on the effects of tagging Lisa J. Natanson John G. Casey Nancy E. Kohler Narragansett Laboratory Northeast Fisheries Science Center National Marine Fisheries Service, NOAA Narragansett, Rhode Island 02882-1199 E-mail address (for L J Natanson) LnatansoawhsunI wh whoi edu Tristram Colket IV 2020 Cordova Ave Vero Beach, Florida 32960 Manuscript accepted 29 September 1998. Fish. Bull. 97:944-953 (1999). Adult tiger sharks, Galeocerdo cuvier, occur worldwide in temper- ate and tropical coastal waters. In the North Atlantic, they reside year round off the coast of Florida and seasonally migrate north as far as Nova Scotia, Canada (Kohler et al., 1995), Additionally, tiger sharks are known to make extensive migra- tions throughout the North Atlan- tic, on occasion traveling to Cuba and Africa,' The tiger shark is listed under the large coastal shark category of the Fisheries Manage- ment Plan for Sharks of the Atlan- tic Ocean (Anonymous, 1993), Al- though it is not a target species of the U,S, inshore longline fishery, small tiger sharks are frequently caught and released alive. Tagging and fishery data indicate that there is a nursery ground for tiger sharks on the continental shelf off the southeast coast of the US,' This area extends from about Augusta, GA, to Daytona, FL, and extends from shore seaward to depths of 100 m, A similar area exists off the coast of North Carolina,- In these areas, tiger sharks of birth size ranging from 61 cm fork length (FL)) to 120 cm FL are commonly caught in the commercial longline fishery. In the Northwest Atlantic, tiger sharks mature between 258 and 265 cm FL (Branstetter et al., 1987) and have been reported to attain a size of 469 cm FL (Castro, 1983), Branstetter et al, ( 1987) used the alternating opaque and translucent bands formed in the vertebral cen- tra to age tiger sharks caught in the western North Atlantic and the Gulf of Mexico, Attempts to verify these estimates with length-frequency analysis proved unsuccessful owing to the inability to distinguish age groups, Branstetter et al, (1987) tried to corroborate the vertebrally derived growth rate with results from one tag-recapture individual; however, this individual's length was estimated at both tagging and recapture. Owing to the high age estimates at L„, Branstetter et al. (1987) suggested that as tiger ' 1992-98. Apex Predators Program, 28 Tarzwell Dr. Narragansett, RI 02882. NMFS unpub. data. ~ Chris Jensen. 1994. Natl. Mar. Fish. Serv, NOAA, 28 Tarzwell Dr, Narragansett, Rl 02882. Personal Commun. Natanson et al : Growth of Galeocerdo cuvier in the western North Atlantic 945 sharks approach maximum sizes, their vertebrae and band deposition may not reflect age. Because at- tempts at verification were unsuc- cessful, and direct validation was not possible, the determination of the tiger shark growth rate was not completely satisfied. To determine if vertebral growth bands reflect age in older individu- als and possibly to verify the esti- mates of Branstetter et al. 1 1987), we undertook a study with tag and recapture data to estimate, inde- pendently, von Bertalanffy param- eters for the tiger shark. Verifica- tion of neonatal and juvenile gi-owth rates was accomplished by using monthly length-frequency data ob- tained over a period of seven years. In addtion, data on growth of oxy tetracycline (OTO- injected tagged and released tiger sharks from an ongoing study were available to compare with length- frequency and tagging growth data. Materials and methods Data from tiger sharks were obtained between 1963 and 1997 from research vessel cruises, sportfishing tournaments, and the commercial shark fishery from Cape Cod, MA, to the Florida east coast. Data for monthly length-frequency analyses were obtained from tiger sharks caught by longline in a delineated area within the nursery grounds off Florida during 1988-94 (Fig. 1). Length measurements Measurements of total length (TL) and FL were taken to the nearest centimeter (cm ) following the conven- tions of Bigelow and Schroeder ( 1948). Fork lengths are reported unless otherwise noted. TL to FL con- versions can be calculated from the relationship Figure 1 Map showing the portion of the nursery area (shaded box) from which monthly length- frequency samples were obtained. cial fishermen who also reported shark size in TL, FL, or weight. All measurements and estimates were converted to FL. Gulland and Holt's ( 1959 ) and Fabens' ( 1965 ) meth- ods were used to calculate von Bertalanffy (1938) growth parameters from the tag-recapture data. Techniques for calculating the parameters according to Gulland and Holt ( 1959 ) came primarily from their publication and additional clarification was obtained from Cailliet et al. (1992). Only fish that were mea- sured at both release and recapture and at liberty for at least 0.9 years were included in the analysis. Two of the three parameters for the von Bertalanffy (1938) growth function (VBGF), k and L„, were esti- mated directly with the methods of Fabens ( 1965 ) and Gulland and Holt ( 1959). T^, cannot be estimated from tagging data alone, rather it requires an estimate of absolute size at age, such as size at birth, and was cal- culated with the VEGF and solving for t,,, such that t.=t ■(l//?)[ln{(L,, L,)IL_ % FL = (TL X 0.8761) 13.3535 r- = 0.99n=44 (Kohleretal., 1995) Li = known length at age (size at birth); k = the von Bertalanffy growth constant; and L . = the theoretical maximum attainable length from the VBGF Tag-recapture data During 1962-96, over 6000 tiger sharks were tagged -with NMFS tags (Casey, 1985) and released as part of the NMFS Cooperative Shark Tagging Program. Tags were returned primarily by sport and commer- The ?Q values were calculated based on an average size at birth of 61 cm FL^ with t = 0. Longevity was estimated from the FL at which >99'^'f of the L. was reached (i.e. 71n2/^) (Fabens, 1965; Cailliet et al., 1992). The von Bertalanffy pa- rameters derived from these methods were compared with growth information obtained from the length- 946 Fishery Bulletin 97(4), 1999 frequency analysis. No OTC-injected individuals were included in these calculations. Tag-recapture with OTC injection During 1985-97, more than 650 tiger sharks (59 to 291 cm FL) were measured, injected with a 25 mg/ kg body weight dose of OTC (Gruber and Stout, 1983), tagged, and released. To determine the effects of OTC on growth, data from recaptured OTC-injected fish were analyzed separately from those of noninjected recaptured fish. Only those OTC-injected specimens measured at both tagging and recapture and at liberty for at least 0.9 years were included in the analysis. For comparison of growth of injected fish to gi'owth of noninjected fish, the growth rates fi-om OTC-injected individuals were plotted with the von Bertalanffy growth function ( VBGF) ft-om the tag-recapture analy- sis. The size at tagging was used as a guide to estimate age at tagging with the VBGF. The time at liberty de- termined the distance along the :*:-axis, and sizes at recapture determined the slope. The growth of the OTC- injected individuals was then compared graphically with the growth curves for long-term (>0.9 yr) tag-re- captured sharks and monthly growth estimates. Monthly growth Data for tiger sharks measured and subsequently tagged from the defined nursery area were analyzed for monthly growth. Data on measured fish were available by month from June 1988 to August 1994. Data were organized into 5-cm intervals. The modes for the 1988, 1989, 1990, 1991, 1992, and 1993 year classes were followed progressively from the birth mode until the last visible mode for that year class. Where modes were not clear (i.e. single fish at more than one interval) the mean was taken as the mode. Previously tagged individuals were not included in this data set. Length-frequency histograms were developed for each month of each year for modal analysis. To determine if the data from the six year classes could be combined, the modes of each year class were plotted by month and compared graphi- cally and through an analysis of covariance. Growth per year was calculated by subtracting the June birth mode from the June one-year mode and the June one- year mode from the June second-year mode. To com- pare the growth of these fish to tagged noninjected fish and tag-recaptured OTC-injected fish, month- per-year growth rates were plotted against the VBGF from the recapture analysis and the growth rates from the individual tag-recaptured OTC-injected fish. The initial positioning of the modes on the jc-axis assumes birth takes place in June^ so that the first point of the monthly growth was fixed on the monthly tag-recapture curve for June. The monthly growth values were fixed on the curve on the basis of month they were calculated for (i.e. June, birth; June, year 1; June, year 2). Graphical comparisons enabled us to determine whether growth based on the tag-re- capture analysis was distinct from growth based on a method without tagging (represented by monthly growth). In addition, we compared Branstetter et al.'s ( 1987) vertebral growth curve to these data. Results Tag-recapture data Information on 42 recaptured tiger sharks, measured at both tagging and recapture and at liberty for at least 0.9 years, was used to produce values of L_ and k of the VBGF (Table D.The Gulland and Holt (1959) method produced the most biologically plausible es- timates of VBGF parameters (Table 2). The Fabens ( 1965 ) analysis underestimated L^ with known maxi- mum size estimates, and the value for k was high. Therefore, further analysis was based on the results from the method of Gulland and Holt ( 1959). Age at maturity, based on lengths at maturity from Branstetter et al. ( 1987), for female (265 cm FL) and male (258 cm FL) tiger sharks is 7 years (Fig. 2; Table 3). Maximum age was estimated to be 27.3 years (335-1- cm FL > 997r of L^). Tag-recapture with OTC injection Analysis of the gi'owth of the four OTC-injected speci- mens recaptured >0.9 years after tagging indicates that individuals grew at appro.ximately the same rate as predicted by the tag-recapture data (Fig. 3). Growth calculated for the first year (growth/year of two individuals tagged at <100 cm FL) was 42.3 and 48.4 cm/yr. and 39.4 and 48.7 cm/yr. for the second year (gi-owth/year two individuals tagged at >100 cm FL) (Table 1). These four individuals were at liberty between 0.94 and 1.19 years. Monthly growth Modes were clearly visible in the length-frequency histograms for small tiger sharks of all year classes from birth to 1.5 years. The modes for 1.5-2 years were less distinct owing to decreased sample sizes at the larger sizes (Fig. 4). Differences in growth rates among the six year classes were statistically signifi- cant (ANCOVA, P>0.05) so we did not combine the monthly length frequencies for all years (Fig. 4). In Natanson et al.: Growth of Galeocerdo cower in the western North Atlantic 947 the first year, the young grew from an average birth size of 65 cm FL to 100-105 cm FL with a growth rate of 40-45 cm/yr. (Fig. 4). More hmited data on second year growth indicated a rate of 35-45 cm/yr Even with this variation, growth in all years paral- leled the tag-recapture growth curve and the indi- vidual growth rates of the OTC specimens (Fig. 3). Discussion It is evident that traditional methods for aging te- leosts do not always work well for sharks. Length- frequency analysis is difficult owing to the slow growth exhibited by most elasmobranch species. Hard part analysis requires validation of the period- icity of band formation, which is often difficult to obtain for shark species. Vertebral age estimates for the sandbar shark, Carcharhinus plumbeus, which was one of the six species considered validated, have been revised since Cailliet's (1990) paper by using tag and recapture evidence (Casey and Natanson, 1992). The new data indicate that although the ver- tebral bands may be formed annually in the young shark, as validated ( Branstetter, 1987b), they are not formed annually throughout the life of the shark and, therefore, band counts severely underestimate age. This type of revision highlights the need for valida- tion of all size classes. It is advisable to use several methods for aging to provide verification for the cho- sen growth curve, particularly if vertebral band counts are used without direct validation. Tag-recapture data can be a useful tool for age and growth determination if accurate measurements are taken at both tagging and recapture and if individu- als are at liberty for a sufficient time for growth to occur. However, problems are associated with this method as well. For example, most length measure- ments are estimated by recreational and commer- cial fishermen. For slow growing sharks, it is impera- tive to obtain accurate length measurements, par- ticularly in large fish. This can prove difficult as well as dangerous at tagging. Therefore, data on large individuals is sometimes lacking and thus will bias results. Additionally, some researchers have shown that tagging with "M" type tags in small sharks, such as the lemon shark, may retard growth (Manire and Gruber, 1991 ). Analysis of the data can also be prob- lematic. The Fabens (1965) method can lead to bi- ased estimates because its basic premise, that tagged individuals are at large for equal time periods, is of- ten violated with sharks. Estimates from that method lead to low values of L^ and high values of ^ (Chien and Condrey, 1987). The Gulland and Holt (1959) method, which allows for unequal times at liberty. Table 1 Tag-recapture data used for growth and growth rate analy- ses. TAL = time at liberty, FLTAG = fork length at tagging, FLCAP = fork length at recapture, 6 = value calculated from Gulland and Holt ( 1959). Also shown are the data for the OTC-injected recaptured fish, although they were not included in growth analyses. Id No. Sex TAL (yr) FLTAG (cm) FLCAP (cm) Average growth/yi (cm) 6 1 M 4.02 75.0 222.0 36.5 0.26 2 F 1.39 76.0 122.0 33.1 0.09 3 M 1.69 76.2 111.8 21.0 0.11 4 F 0.90 77.0 124.0 52.3 0.06 5 F 1.42 78.7 114.3 25.1 0.09 6 F 1.75 79.0 136.0 32.5 0.12 7 M 0.97 85.0 130.0 46.6 0.06 8 F 1.60 86.0 171.0 53.0 0.11 9 F 1.22 88.0 128.0 32.8 0.08 10 F 0.91 89.0 120.0 34.0 0.06 11 F 0.92 90.0 155.0 70.9 0.06 12 F 4.90 91.0 211.0 24.5 0.32 13 M 1.00 91.0 122.0 30.9 0.07 14 F 4.90 91.0 211.0 24.5 0.32 15 M 2.04 91.0 168.0 37.7 0.13 16 F 1.03 94.0 136.5 41.2 0.07 17 F 1.60 96.0 157.0 38.1 0.11 18 M 1.01 97.0 147.0 49.5 0.07 19 F 1.51 98.0 154.0 37.0 0.10 20 M 0.97 98.0 142.0 45.4 0.06 21 F 3.28 98.0 195.0 29.6 0.22 22 F 0.95 102.0 157.0 57.9 0.06 23 F 0.90 104.0 124.0 22.1 0.06 24 M 0.99 106.0 156.0 50.4 0.07 25 M 1.00 106.0 160.0 54.2 0.07 26 M 1.37 108.0 147.0 28.4 0.09 27 F 0.97 109.0 160.0 52.6 0.06 28 F 1.07 110.0 173.5 59.6 0.07 29 M 0.94 111.0 156.0 47.9 0.06 30 F 1.56 114.0 155.0 26.2 0.10 31 F 2.20 117.0 184.0 30.5 0.14 32 F 1.69 117.5 175.5 34.3 0.11 33 M 0.90 118.0 152.0 37.7 0.06 34 M 1.48 120.0 167.0 31.8 0.10 35 M 0.91 121.0 161.0 43.9 0.06 36 F 1.37 124.0 173.0 35.7 0.09 37 M 0.91 124.0 137.0 14.2 0.06 38 F 0.93 149.0 164.0 16.2 0.06 39 M 1.16 170.0 193.6 20.3 0.08 40 F 5.28 190.0 235.0 8.5 0.35 41 F 1.20 281.0 288.0 5.8 0.08 42 M 3.33 287.0 317.5 9.2 0.22 218097 F 0.94 104,0 141.0 39.4 304303 M 0.98 70.5 112.0 42.3 190804 F 1.19 100.0 158.0 48.7 204379 M 1.06 93.5 145.0 48.4 948 Fishery Bulletin 97(4), 1999 350 -T This Study ^ "^^Z^ " " 300 - ^ ^^' ^^^^ .^ -^^ ^^ Branaener er a/ (1987) ^ M 250 - / y , . / • I 200 / y c / y / / p ? / / S o / / |l50i / / / / a LL / / // s v // Si c 100 ■ / c 5 § m // s f a CO 50 T S £ OQ >> ■■E % 5 , . , u 1 ! i , . i . 1 0 2 4 6 8 10 12 14 16 18 20 Age (years) Figure 2 Tag-recapture growth curve derived in this study by using the Gulland and Holt ( 1959) method compared with the growth curve derived by Branstetter et al.'s ( 1987 ) using Atlantic Ocean vertebral data. Estimated size and age at maturity are included on both curves Table 2 VBGF parameters calculated by using two different tag-recapture methods (Gulland and Holt, 1959, and Fabens, 1965) (n=\1\ and compared with Branstetter et al.'s ( 1987 ) estimates derived from vertebral analysis ( fork lengths calculated by using Branstetter et al.'s ( 1987) conversions). Model Gulland and Holt, 1959 Fabens, 1965 Branstetter et al, 1987 Atlantic Gulf of Mexico 337 FL 293 FL 440 TL ( 365 FL) 388 TL (324 FL) SE 6.95 19.10 0.107 0.184 0.178 0.217 -2.35 -1.13 SE 0.048 0.031 -1.12 -1.08 therefore, appears to be more appropriate for sharks (CailUetetal., 1992). In this study, the Gullanci and Holt ( 1959) method produced more biologically reasonable results than the Fabens (1965) method. The L,^ calculated from the Gulland and Holt method was 337 cm FL, which is lower than the maximum reported western North Atlantic values from the literature (391 cm FL; Bigelow and Schroeder, 1948). More recently, in the western North Atlantic and Gulf of Mexico, the larg- est sharks observed were 346 cm FL (Branstetter, 1981) and 339 cm FL^ the latter quite close to our ^ Kohler, N. E., H. W Pratt Jr. L. J. Natanson, P. Turner, and R. Briggs. 1996. The shark tagger 1995 summary Narragan- sett Laboratory, Northeast Fish. Science Center, Natl. Mar Fish. Serv., NOAA, 16 p. Natanson et al.: Growth of Galeocerdo cuvier m the western North Atlantic 949 estimate. However, the tiger shark has been reported to 469 cm FL (Castro, 1983 ). The L„ value estimated from the Fabens (1956) method (293 cm FL), how- ever, is lower than all reported values. In addition, the Fabens (1965) method value for k is high (Table 2). We therefore concluded that the Gulland and Holt (1959) VBGF is the more appropriate model to use for the tiger shark. Cailliet et al. (1992) and Van Dykhuizen and Mollet (1992) also preferred the Gulland and Holt (1959) model over that of Fabens (1965) for the angel and sevengill sharks, respectively. Data from this study indicated that neither tagging nor tagging combined with OTC injection appears to retard the growth rate in neonate and juvenile sharks up to 150 cm FL (Fig. 3). To the contrary, the growth rates from tagged sharks were higher than estimates obtained from vertebral gi-owth bands (Table 3). The growth rates for the first two years of life for the tiger shark estimated from monthly length frequencies, tag- recapture (Gulland and Holt. 1959), and tag-recapture with OTC were all similar (Fig. 3). Tanaka ( 1990) found that growth rates of OTC-injected Japanese wobbe- gongs, Orectolobusjaponicits, and neonate swell sharks, Cephaloscyllium umbratile, were not significantly dif- ferent from controls. More recently, Gelsleichter et al. ( 1998 ) evaluated the toxicity of OTC on gi-owth rates of captive nurse sharks, Ginglymostoma cirratum, and concluded that there were no adverse effects of OTC on growth rate. These data support the use of OTC as an effective method for determining vertebral band periodicity without inten-upting normal growth pat- terns (Tanaka, 1990; Gelsleichter et al., 1998). Our data do not verify nor refute age estimates for the tiger shark previously obtained from vertebral band counts (Branstetter et al., 1987 )( Table 2; Fig. 2). The k value from our study is higher than Branstetter et al.'s (1987) Atlantic value and our L is lower These differ- ences are to be expected when comparing VBGF pa- rameters from a conventional age-length (vertebral methods) study with those derived from a growth in- crement (tagging) study (Sainsbury, 1980; Francis, 1988). These types of curves are not directly compa- rable because the parameters are derived differently and, therefore, have different meanings (Francis, 1988). However, comparison of the estimates obtained in these studies to known values, such as size at birth and maxi- mum size, can provide insight into the fit of the curves. This information allows us to determine which curve is best suited to be used for age at maturity and maxi- mum age estimates. We believe that the current tag- recapture values are more accurate on the basis of veri- fication available from the monthly length-frequency analysis and the consistency of the estimates to mea- surable parameters such as size at birth and maximum size. The L^ calculated by Branstetter et al. ( 1987) is Table 3 Size at age and growth per year for the tiger shark, Gateo- cerdo cuvier, calculated for tag and recapture data with | Gullan d and Holt's (1959) (this study) method compared with B -ans tetter et al.'s ( 1987 ) Atlantic vertebral data. Ap- proximate size at matu rity is indicated by bold typeface. Age Size (cm, FL) Branstetter' This study (years This study 1987 growth/yr (cm) Birth 61 73 1 106 103 45 2 144 130 38 3 175 153 32 4 202 175 26 5 224 194 22 6 242 212 18 7 258 227 15 8 271 241 13 9 281 254 11 10 290 265 9 11 298 275 8 12 304 284 6 13 310 293 5 14 314 300 4 1.5 318 307 4 16 321 313 3 17 324 318 3 18 326 323 2 19 328 327 2 20 329 331 2 ' Brans tetter et al. (1987 values converted oFL. lower than maximum reported sizes, and the size at birth (73 cm FL), based on the von Bertalanffy curve from that study, was high as related to known param- eters (60-65 cm FL^ ). Statistically significant differences between age es- timates from the Gulf of Mexico and North Atlantic populations of tiger shark found by Branstetter et al. (1987) may not be biologically significant (Yoccoz, 1991). The differences in age at maturity obtained between these two areas is only 2-3 years. Consid- ering the relatively slow growth and large overlap of size at age for this species, a 2-3 year difference could be included in the realm of measurement error. In addition, we found statistically significant differences in growth by year in first year tiger sharks obtained from the same region over a period of five years. The differences in growth between years indicate that tiger shark growth is quite variable and probably dependent on fluctuations of many parameters in- 950 Fishery Bulletin 97(4). 1999 •Id (1Se7) TMi S&idy. Qulwd aod Holt (1958) HMhod IMofittily LangSi Fr«qu«ncy Modes. 6 y««n OTC. 4 ndlvWuaO 10 15 20 25 30 35 Figure 3 The von Bertalanffy growth curves from tag-recapture analysis was estimated with the Gulland and Holt ( 1959) model and the von Bertalanffy growth curve estimated from vertebral analysis (Branstetter et al. 1987). The solid lines represent the individual monthly length-frequency mode data from six year classes (1988-93). The initial points of the line are set by the size in June of the birth year The solid lines are actual growth of four individual OTC-injected recaptured sharks at liberty over 0.9 yr. The length at tagging was used to obtain the initial age estimate to set the point on the graph. eluding environmental conditions and prey availabil- ity. Our data show that that the majority of neonatal tiger sharks remain in the nursery area from birth until about 120-150 cm FL or 1.5—2 years of age. ^ It is known that tiger sharks frequently migrate into and out of the Gulf of Mexico (Kohler et al., in press). It is doubtful, with the migratory nature of this spe- cies and mixing between these areas, that growth rate differences in these groups are biologically sig- nificant past perhaps the first three years of life. The longevity of the tiger shark is difficult to esti- mate. The NMFS tagging program has received data on five tiger sharks at liberty for 6 to 11 years. The oldest of these fish would have been 3+ years at tag- ging (185 cm FL, estimated), on the basis of the Gulland and Holt (1959) growth curve and, there- fore, 14-1- years at recapture (325 cm FL, measured). Branstetter et al.'s ( 1987 ) oldest aged tiger shark was 16 years of age. Our longevity estimate, based on a 7 half-life criterion, indicated that tiger sharks may live to be at least 27 years of age. Branstetter et al. (1987) estimated maximum age at anywhere from 20-37 years based on L . for their various VBGF curves and rate of growth of large individuals. Based on revised growth estimates presented in this study, estimated maximum age for this species is 27 years. Age at maturity, estimated from tag-re- capture data from this study and size at maturity estimates from Branstetter et al. (1987), is seven years suggesting that females mature at 25% of their maximum age and may reproduce 10 times based on a two-year reproductive cycle. Overall, the tiger shark is similar to other large carcharhinids in that it grows slowly and has a rela- tively long life, although it matures earlier than many other species in the northwest Atlantic (C. leucas 14— 18 years, Branstetter and Stiles, 1987; C. plumbeus 15-30 years; Casey and Natanson, 1992, Sminkey and Musick, 1995; C. obscurus 19-20 years; Natan- son, 1994; C. falciformis 6.5-12 years, Branstetter, 1987c, Bonfil et al., 1993). Branstetter (1987b) dis- cussed the various life strategies of sharks on the basis of their k values. He suggested that tiger sharks, with k values from 0.11 to 0.16, fit into an intermediate category between slow growth species (/?=0.05-0.10) such as C. plumbeus, C. obscurus, Neg- aprion brevirostris, and Sphyrna lewini and fast growth species (/f>0.2) such as C. limbatus, C. brevipinna, Rhizoprionodon terraenovae, and Prionace glauca. Other sharks in the intermediate group include C. acronotus and C. falciformis according to Branstetter ( 1987b). The tiger shark reaches maturity at a lower percent {257c) of its total age than do other carcharhinids for which this parameter has been es- timated (S. lewini 33-50^^r, Branstetter, 1987c). This finding suggests that the tiger shark has a longer re- productive life span and possibly a greater reproduc- tive potential than other carcharhinids. Natanson et al : Growth of Galeocerdo cuvierm the western North Atlantic 951 25 20 15 10 5 0 June 1992 n =61 1 Year Birth Mode ^105 cm FL Mode = 65cm FL ■■! ■Illll February 1993 n =61 25 20 15 10 5 0 25 20 15 10 5 0 July 1992 rt, = 31 8 8 2 8 S S R S 8 August 1992h =33 8 8 £ 8 S ? S September 1991 n = 21 En 35 r-- 35 Cfi o August 1993 n 25 T 20 I 15 I 10 i 5 0 ° § ? ? '? = 116 ; .ll. ..iillli. 88Sga82SS9S Fork length (cm) Fork length (cm) Figure 4 Monthly length-frequency data for the 1992 year class used for obtaining length-frequency growth rates for small (<160 cm FL) tiger sharks in the nursery area. First and partial second year growth for the 1992 year class is represented with the solid line. Second and partial third year growth for the 1991 year class is represented by the dashed line. 952 Fishery Bulletin 97(4), 1999 Casey and Natanson (1992) showed the advantage of using tag-recapture data over vertebral analysis for aging the sandbar shark. Cailliet et al. (1986) reviewed the techniques available for determining age and verifying age estimates in elasmobranchs. They pointed out that ages estimated from growth zones in calcified hard parts need to be verified with other methods, such as length-frequency and tag- recapture analyses. The authors also stressed the importance of validating the temporal periodicity of the calcified bands with tag-recapture data from the laboratory or field, coupled with OTC marking. Cailliet ( 1990), in an update of the Cailliet et al. (1986) review, listed the studies, to that date that had employed the various verification methods. Sev- eral of these used the combination of tag-recapture, length-frequency and OTC marking for verification or validation (or both) of the calcified structure age estimates. Growth estimates from a laboratory and field study of the Atlantic sharpnose shark [Rhizo- prionodon terraenovae) (Branstetter 1987a) corre- sponded well to estimates generated with length fre- quencies and vertebral rings (Parsons, 1985). Casey et al. ( 1985 ) used length-frequency, vertebral and tag- recapture analyses to verify age estimates for the sandbar shark, C. phimbeiis, and concluded that there was close agreement with all three methods. Pratt and Casey (1983) also used these three meth- ods to age the shortfin mako, Isurus oxyrinchus, and concluded that not only was there agreement between the methods but that tagging did not affect growth in this species. Smith (1984) validated the periodic- ity of vertebral band deposition in the leopard shark, Triakis semifasciata using OTC. Kusher et al. ( 1992 ) were able to confirm these results as well as use length-frequency data and additional tag-recapture data to produce independent age estimates for this species. The tag-recapture curve gave slower k val- ues than the vertebral, and although the k values were not significantly different, the authors sug- gested that tagging may have had an effect on growth in this species (Kusher et al. 1992). These results accentuate the requirement of good tag and recap- ture data as a backup for vertebral studies, particu- larly if combined with OTC injections for validation. In the case of the tiger shark in this study, tag-re- capture and length-frequency data have provided independent estimates of growth for verification. In addition, the results present evidence against the suggestion that tagging, with or without OTC injec- tion, decreases the growth rate of sharks. It can be argued that the tiger shark cannot be used to gener- alize about sharks because they grow rapidly in re- lation to many other species. A birth size of 61 cm FL with a corresponding weight of 1.8 kgs. is small in relation to many large coastal species, such as the dusky shark ( size and weight at birth: 81 cm FL and 7 kgs., respectively Castro, 1983; Kohler et al., 1995) and the silky shark (size and weight at birth: 64 cm FL and 9 kgs., respectively). If these relatively small tiger shark young can withstand the rigors of tag- ging and continue to grow at a similar rate as untagged individuals, then it is certainly reasonable to believe that a larger shark can as well. Regard- less, all species need to be evaluated individually for their reactions to tagging and OTC injection. Acknowledgments We are thankful to the thousands of fishermen who voluntarily tag and return tags to us and thus make this continuing program possible. We thank Chris Jensen who helped in tagging and injecting tiger sharks with tetracycline as well as providing accu- rate measurements on both tagged and recaptured specimens. Fisherman Eric Sander provided dozens of vertebrae from precisely measured recaptured young tiger sharks as well as from nontagged speci- mens. We also thank Steve Branstetter for discus- sions on the phone and for reviewing the manuscript. Greg Cailliet and Sabine Wintner also provided in- valuable comments on the manuscript. We appreci- ate the encouragement and assistance of our col- leagues in the Apex Predators Program. Literature cited Anonymous. 1993. Fishery management plan for sharks of the Atlantic Ocean. U.S. Dep. Commer. NOAA, NMFS, Silver Spring, MD. 167 p. Bigelow, H. B., and W. C. Schroeder. 1948. Shark.s. In J. Tee-Van. C. M. Breder. S. F. Hilde- brand, A. E. Parr, and W. C. Schroeder (eds. ). Fishes of the Western North Atlantic, part one, vol. 1, p. 59-546. Mem. Sears Found. Mar. Res., Yale Univ. Bonfil, R., R. Mena, and D. de Anda. 1993. Biological parameters of commercially exploited silky sharks. Carcharhmus falciformis, from the Campeche Bank, Me.\ico. /;; S. Branstetter (ed. ). Conservation biol- ogy of elasmobranchs, p. 73-86. U.S. Dep. Commer, NOAA Tech Rep. NMFS 115. Branstetter, S. 1981. Biological notes on the sharks of the North Central Gulf of Mexico. Contrib. Mar. Sci. 24:13-34. 1987a. Age and growth validation of newborn sharks held in laboratory aquaria, with comments on the life history of the Atlantic sharpnose shark, Rhizoprionodon terraenovae. Copeia. 1987(21:291-300. 1987b. Age and growth estimates for blacktip, Carcharhmus limhatus, and spinner, C. brevipmna, sharks from the north- western Gulf of Mexico. Copeia. 1987(4): 964-974. Natanson et al.: Growth of Galeocerdo cower in the western North Atlantic 953 1987c. Age, growth and reproductive biolog>' of the silky shark, Carcharhmus falciformis. and the scalloped ham- merhead, Sphyrna lewini. from the northwestern Gulf of Mexico. Environ. Biol, of Fish. 19(3):161-173. Branstetter, S., J. A. Musick, and J. A. Colvocoresses. 1987. A comparison of the age and growth of the tiger shark, Galeocerdo curieri. from off Virginia and from the North- western Gulf of Mexico. Fish. Bull. 85(2):269-279. Branstetter, S., and R. Stiles 1987. Age and growth estimates of the bull shark, Carcha- rhmus leiicas, from the northern Gulf of Mexico. Environ. Biol. Fish. 20(3):169-181. Cailliet, G. M. 1990. Elasmobranch age determination and verification: an updated review. In H. L. Pratt Jr., S, H. Gruber, and T. Taniuchi (eds.), Elasmobranchs as living resources: ad- vances in the biology, ecology, systematics, and status of the Fisheries, p. 157-165. U.S. Dep. Commer., NOAA Tech. Rep. NMFS 90. Cailliet, G. M., H. F. Mollet, G. G. Pittenger, D. Bedford, and L. J. Natanson. 1992. Growth and demography of the Pacific angel shark [Squatina californica). based upon tag returns off Cali- fornia. Aust. .J. Mar. Freshwater Res. 43:1313-1330. Cailliet, G. M., R. L. Radtke, and B. A. Welden. 1986. Elasmobranch age determination and verification: a review. In T. Uyeno, R. Aral, T. Taniuchi, and K. Matsuura (eds.), Indo-Pacific fish biology: proceedings of the second international conference on Indo-Pacific fishes, p. 345- 360. Ichthyological Soc. of Japan, Tokyo. Casey, J. G. 1985. Transatlantic migrations of the blue shark: a case history of cooperative shark tagging. In R. H. Stroud (ed.). World angling resources and challenges: proceedings of the first world angling conference. Cap d'Agde, France, Sep- tember 12-18, 1984, p. 253-268. Int. Game Fish Assoc. Ft. Lauderdale, FL. Casey, J. G., and L. J. Natanson. 1992. Revised estimates of age and growth of the sandbar shark tCarcharhinus plumbeus) from the western North Atlantic. Can. J. Fish. Aquat. Sci. 49(7):1474-1477. Casey, J. G., H. L. Pratt, and C. E. Stillwell. 1985. Age and growth of the sandbar shark iCarcharhmus plumbeus) from the western North Atlantic. Can. J. Fish. Aquat. Sci. 42:96.3-975. Castro, J. I. 1983. The sharks of North American waters. Texas A&M Univ Press, College Station, TX, 180 p. Chien, Y-H , and R. E. Condrey. 1987. Bias in estimating growth parameters using Fabens" mark-recapture procedure. Asian Fish. Sci. 1( 1987 1:65-74. Fabens, A. J. 1965. Properties and fitting of the von Bertalanffy growth cur\-e. Growth. 29:26.5-289. Francis, R. I. C. C. 1988. Are growth parameters estimated from tagging and age-length data comparable? Can. J. Fish. Aquat. Sci. 45:936-942. Gelsleichter, J., E. Cortes, C. A. Manire, R. E. Hueter, and J. A. Musick. 1998. Evaluation of toxicity of oxytetracycline on growth of captive nurse sharks, Gmglymostoma cirratunj. Fish. Bull. 96:624-627. Gruber, S. H., and R. G. Stout. 1983. Biological materials for the study of age and growth in a tropical marine elasmobranch, the lemon shark, Negaprion brevirostns (Poeyl. In E. D. Prince and L. M. Pulos (eds. I. Proceedings of the international workshop on age determination of oceanic pelagic fishes: tunas, billfishes, and sharks, p. 193-205. U.S. Dep. Commer., NOAA Tech. Rep. NMFS 8. Gulland, J. A., and S. J. Holt. 1959. Estimation of growth parameters for data at unequal time intervals. Journal du Conseil. Cons. Int. pour I'Explor. de la Mer 25:47-49. Kohler, N. E., J. G. Casey, and P. A. Turner. 1995. Length-weight relationships for 13 species of sharks from the Western North Atlantic. Fish Bull. 93(21:411- 417. In press. NMFS Cooperative Shark Tagging Program, 1962-1993: an atlas of shark tag and recapture data. U.S. Dep. Commer, NOAA Tech. Mem. NMFS-F/NEC. Kusher, D. I., S. Smith, and G. M. Cailliet. 1992. Validated age and growth of the leopard shark, Tnakts semifasciata. with comments on reproduction. Environ. Biol. Fish. 35:187-203. Manire, C. A., and S. H. Gruber. 1991. Effect of M-type dart tags on field growth of juvenile lemon sharks. Trans. Am. Fish. Soc. 120:776-780. Natanson, L. J. 1994. Age and growth estimates for the dusky shark, Carcharhinus obscurus. in the western North Atlantic. Fish. Bull. 93: 116-126. Parsons, G. R. 1985. Growth and age estimation of the Atlantic sharpnose shark, Rhizoprionodon terraenovae: a comparison of techniques. Copeia 198.5( l):80-85. Pratt, H. L., and J. G. Casey. 1983. Age and growth of the shortfin mako, Isurus oxyrinchus. using four methods. Can. J. Fish. Aquat. Sci., 40(11): 1944-1957. Sainsbury, K. J. 1980. Effect of individual variability on the von Bertalanffy growth equation. Can. J. Fish. Aquat. Sci. 37:241-247. Sminkey, T. R., and J. A. Musick. 1995. Age and growth of the sandbar shark, Carcharhinus piunjbeus. before and after population depletion. Copeia 1995(4):871-883. Smith, S. E. 1984. Timing of vertebral-band deposition in tetracycline- injected leopard sharks. Trans. Am. Fish. Soc. 113:308- 313. Tanaka, S. 1990. Age and growth studies on the calcified structures of newborn sharks in laboratory aquaria using tetracycline. In H. L. Pratt Jr. S. H. Gruber, and T. Taniuchi. (eds.), Elasmobranchs as living resources: advances in the biol- ogy, ecology, systematics, and status of the fisheries. U.S. Dep. Commer., NOAA Tech. Rep. NMFS 90:189-202. Van Dykhuizen, G., and H. F. Mollet. 1992. Growth, age estimation and feeding of captive sevengill sharks, Notoryhchus cepedianus. at the Monterey Bay Aquarium. Aust. J. Mar Freshwater Res. 43:297-318. von Bertalanffy, L. 1938. A quantitative theory of organic growth (inquiries on growth laws II). Hum. Biol. 10:181-213. Yoccoz, R. G. 1991. LIse, overuse, and misuse of significacne tests in evo- lutionary biology and ecology. Bull. Ecol. Soc. Am. 72(2): 106-111. 954 Abstract.— Understanding the rela- tive importance of pre- and postsettle- ment processes is critical to under- standing the population dynamics of marine fishes. Our goals in this study were 1 1 to examine habitat preference and habitat use of newly settled Atlan- tic croaker, Micropogonias undulatus, and 2) to determine if postsettlement growth or predation varied with habi- tat type. Field surveys showed no dif- ference in croaker abundance among three estuarine habitats: marsh edge, seagrass, and sand. Behavioral e.xperi- ments in laboratory mesocosms sug- gested that the pattern of similar use of habitats in the field results from a lack of preference among habitats. In a field experiment, croaker recruitment was greater to artificial seagrass than to sand habitats, but there was no dif- ference in fish density in habitats with or without food supplementation. More- over, growth rates were similar in both sand and artificial seagrass habitats and in habitats with or without food supplementation. In a second experi- ment, we were unable to detect a dif- ference in the density of newly settled croaker between sand and artificial seagrass habitats, or between habitats with predator access limited by cages and cage controls. Our results demonstrate that newly settled croaker use different estuarine habitats similarly, and there does not appear to be a fitness conse- quence of using many habitats. We sug- gest that for habitat generalists, such as the Atlantic croaker, variability in larval supply will be a stronger predictor of population dynamics than will variabil- itv of habitat attributes. Recruitment of Atlantic croaker, Micropogonias undulatus: Do postsettlement processes disrupt or reinforce initial patterns of settlement?* Rachel Petrik Phillip S. Levin Institute ol Marine Science University of California, Santa Cruz Santa Cruz, California 95064, E-mail address (for P S Levin, contact author) levinraJbiology ucsc edu Gregory W. Stunz Department of Marine Biology Texas A&M University, Galveston, Texas 77553 John Malone Department of Biology University of California, Los Angeles, California 90095 Manuscript accepted 21 .^pril 1999, Fish. Hull. 97:9,'')4-961 1 1999). Understanding the causes of fluc- tuations in population abundance is critical for ecologists and fishery biologists. For marine fishes with life histories in which adults have limited home ranges and larvae are pelagic and advected vast distances from natal sites, an understanding of variability in larval supply to local populations is critical for tmderstand- ing the mechanisms that produce dy- namics in populations (Caley et al., 1996). In addition, habitat selection by settling fish (Carr, 1991; Levin, 1991; Wellington, 1992; Tolimieri, 1995), and habitat-specific growth and mortality (Heck and Orth, 1980; Hixon and Beets, 1993; Levin et al., 1997 ) may ultimately reinforce or dis- loipt patterns created by variable lar- val supply (Jones, 1997). Thus, knowledge of the degree to which pro- cesses such as habitat selection, com- petition, or predation modify initial patterns of lai'val settlement is im- portant in understanding the popu- lation dynamics of marine species. The importance of variability in postsettlement growth or mortality and the level to which postsettlement processes alter initial patterns of larval settlement can be a function of habitat structure. For example, on coral reefs, holes provide a ref- uge from predation, and on reefs with large numbers of holes, the importance of predation is reduced (Shulman, 1984; Hixon and Beets, 1993). Similarly, Atlantic cod settle in equivalent densities in a variety of habitats but suffer lower preda- tion rates in structurally complex habitats (Tupper and Boutilier, 1995). Thus, habitat-specific mor- tality disrupts initial patterns of larval settlement. Differences in habitat structtu'e may also impact growth rates or the ability offish to procure food (Nelson, 1979; Heck and Thoman, 1981; Stoner, 1982). As examples, 1 ) pinfish have greater success capturing amphipods in shoal grass iHalodule wrightii) than in similar densities of turtle grass ( Thalassia te^tudin iiin ) ( Stoner, ' Contribution 10 of the Partnership for Inter- disciplinary Studies of Coastal Oceans I PISCO I: a long-term ecological consortium funded by the David and Lucile Packard Fiiundation. Petrik et al.: Recruitment of Mkropogonias undulatus 955 1982); 2) Atlantic cod grow faster in seagi-ass habitats than in sand, rocky reef, or cobble habitats (Tupper and Boutilier, 1995); and 3) pinfish exhibit higher growth rates in seagrass than in sand habitats (Levin etal., 1997). Atlantic croaker, Micropogonias undulatus (here- after referred to as croaker), range from Cape Cod to Campeche Bank, Mexico (Johnson, 1978), and occur both offshore and in estuaries in a variety of habi- tats including mud, sand, and seagrass (White and Chittenden, 1977; Johnson, 1978; Rooker et al., 1998). Croaker are an important component of com- mercial fisheries in the Gulf of Mexico and south- eastern United States, often dominating bottom fish landings, and are an important sport fisheiy in this region (Lassuy, 1983). In the Gulf of Mexico, croaker spawn over the continental shelf or near inlets from September to May with peak levels occurring before January (Johnson, 1978; Cowan, 1988; Cowan and Shaw, 1988). Larval croaker then move toward shore and may be transported hundreds of kilometers be- fore entering estuarine nursery grounds (Cowan and Shaw, 1988; Norcross, 1991). In Texas, recmitment of croaker peaks in November (Rooker et al., 1998). It is not clear whether variability in abundance of juvenile croaker is the result of variability in lai-val supply or differential postsettlement growth and mortality. The delivery of larval croaker recruits to estua- rine nursery habitats is dependent on large-scale oceanographic processes (Cowan and Shaw, 1988). Once fish arrive at estuaries, delivery into suitable habitats is dependent on currents and tidal processes (Norcross, 1991 ). As fish that are competent to settle approach nursery grounds they have the opportu- nity to choose specific microhabitats. In this paper we examined patterns of microhabitat preference and use by newly settled croaker, as well as the conse- quences of microhabitat associations. Specifically we asked 1) Do croaker have specific microhabitat pref- erences and are these preferences reflected in pat- terns of abundance in the field? 2) Does food supply limit the number or gi'owth rates of croaker recruits in different habitats? 3) Does predation determine the number of recruits in different habitats? Methods Habitat use by newly recruited croaker To determine what habitats newly recruited Atlan- tic croaker use, we conducted a field sui-vey during November 1996 at Christmas Bay (29"03'N, 95'10'W), near Galveston, TX. Christmas Bay is a shallow estu- ary and contains the most easterly well-developed seagrass bed in Texas. A detailed description of this site can be found in Thomas et al. (1990). The seagi-ass bed is dominated hy Halodule wrightii with an average density of 10,469 shoots/m^ (SE=461). An epibenthic sled was used to quantify fish abundance in three habitats: bare sand, seagrass meadow, and marsh edge. We defined marsh edge habitat as the subtidal substrata directly adjacent to a Spartina alterniflora marsh. The sled consisted of a 0.66 m x 0.5 m opening fitted with a3-mlongnet(l-mmmesh) with a removable codend. Habitats were sampled by placing the sled on the substratum, extending a 15-m rope in a semicircular fashion (to avoid disturbing sampling area) and pulling the sled through a lO-m^ area. Each habitat was sampled four times at two different sites, resulting in eight samples per habi- tat type. Differences in croaker density were exam- ined with a two-way analysis of variance with both site and habitat type as fixed effects. In this and sub- sequent analyses, if we failed to reject the null hy- pothesis of no difference in croaker abundance be- tween habitats, then power analysis was performed. If statistical power was low, we calculated the num- ber of replicates required to achieve sufficient power to accept the null hypothesis. To examine habitat preference we performed choice experiments in laboratory mesocosms. Six 117-L mesocosms were constructed from round circular plastic tanks (41.3 cm diameter x 60 cm). The mesocosms were filled with 5 cm of sand, a plastic mesh screen was placed on top of the sand, and an additional 5 cm of sand was placed over the mesh. Each tank was filled with filtered seawater and main- tained at ambient light and temperatures. We divided mesocosms in half, with each half randomly receiv- ing a sand or grass habitat. Sand habitat was the sand bottom described above. To construct seagrass habitats, cores of seagrass were randomly collected from the field and brought to the laboratory where they were washed and dipped in fresh water. After leaves were wiped to remove any epiphytic growth, the cores were planted in each mesocosm. One croaker ( 15-20 mm SL) was introduced to the center of each mesocosm and monitored for any ab- normal behavior for 24 h. After the initial acclima- tion period, the location of each croaker was visually determined hourly for ten consecutive hours. Visual observations were performed by a single observer peering into the mesocosm, without disturbing the fish. This was repeated for six mesocosms over two days for a total of 12 mesocosm observations. New fish were used for each trial. Percent occurrence in each habitat was determined for all twelve trials. A one-way Ntest determined if percent occurrence in seagi'ass was different from 509f . 956 Fishery Bulletin 97(4), 1999 Effects of food supply on recruitment and growth of croaker in varying habitats Field experiments were conducted in East Lagoon, located at the eastern most end of Galveston Island, TX (29°20'N, 94°44'W). East Lagoon is 1.6 km long, 0.48 km wide, and has a maximum depth of 4.6 m. Water is exchanged tidally by means of seven 0.92 m- diameter cement culverts with the Galveston Ship Channel, which runs from the Gulf of Mexico into Galveston Bay. A detailed description of this site can be found in Levin et al. ( 1997). Seagrass, once wide- spread in Galveston Bay, including East Lagoon, has decreased by 90% from peak levels (Pulich and White, 1991); no natural seagrass habitats presently are found in East Lagoon. The absence of natural seagrass beds allowed us to establish artificial seagrass beds with desired characteristics, without the confounding effects of a natural seagrass bed. Experiments were located >8 m from the Sparfina alterniflora dominated marsh edge and placed at an average low tide depth of 42 cm. To test the null hypothesis that food supply does not limit abundance or growth rates of croaker re- cruits in different habitats, we conducted an experi- ment in which food supply was manipulated in sand and seagrass habitats. On 20 February 1996 we cre- ated five blocks each consisting of four 1-m- experi- mental plots. Within each block, food supply and habitat type were manipulated orthogonally. To con- trol for differences in seagi'ass structure or seagrass- associated food resources, we used artificial seagrass habitats. Artificial seagrass habitats were con- structed from a 1-m- polyvinylchoride (PVC) ( 1.3 cm diameter) frame, strung with monofilament to form a grid consisting of 576 points. At each of these points a 16 cm X .5 cm strand of green ribbon was woven in, such that the frame consisted of 576 shoots of seagrass, each shoot having two leaves. No exces- sive fouling was observed on the frame or ribbon for the duration of the experiment. We performed a pre- liminary experiment to determine if the structure of the PVC-frame would attract more fish recruits than bare sand, and no difference was found between the bare sand plot and the PVC-frame (F, 2,3=0.512, 0!=0.61, 1-/3=0.76). Consequently we performed sub- sequent experiments without a PVC-frame control. Food supply was experimentally manipulated with feeding tubes in each experimental plot. Feeding tubes were constructed of a 7.5-cm diameter length of PVC pipe attached to a l.S-cm diameter PVC pipe stake, with the bottom of the tube about 15 cm from the substratum, and the top always above the water line. We provided supplemental food daily for seven days, from 23 February to 3 March 1996, to half of the sand and seagrass replicates (i.e. five sand and five seagrass plots received food). Food consisted of 200 g of fish fiesh and 300 mL of water blended to produce plankton-size particles (Forrester, 1990; Levin et al., 1997). The fish puree was placed in ice cube trays and frozen. Each frozen cube yielded 11.8 g offish flesh. One cube of frozen food was placed in the feeding tubes of appropriate replicates, whereas control plots received one ice cube and no food was added. As the ice cube containing food melted, it delivered a continuous stream of particles to the habitat for 5-15 min. We observed fish readily consuming the supplemented food in both the field and laboratory. On 4 March 1996, the experiment was terminated by sampling each plot. Recruit density was quanti- fied by using 1 m'' (1x1x1 m) drop samplers (Zimmerman et al., 1984; Fonseca et al., 1990). Drop samplers were constructed of 9.5-mm diameter rebar covered on four sides with taut 2-mm nylon mesh. A dip net (90 x 100 cm, 2-mm nylon mesh) was used to retrieve fish from the samplers, and replicates were considered adequately sampled when five consecu- tive passes of the dip net yielded no fish (Fonseca et al., 1990). A blocked two-factor analysis of variance was used to test the hypotheses that the abundance of newly recruited croaker did not vary among habi- tat or food supplemented treatments. Five fish from each replicate were haphazardly selected for further analysis. We measured the 80 selected fish to the nearest 0.1 mm (SL), removed their otoliths, and stored them in immersion oil for one week. Fish age was then determined by enumer- ating the daily growth rings on the lapillar otolith by using an image analysis system. The existence of daily rings on croaker otoliths has been validated previously ( Nixon and Jones, 1997 ). Each otolith was examined independently three times. If two of the three counts did not agree, the fish was discarded and another selected. When two of the three counts were the same, that count was used as a datum in the analysis. Differences in growth rates were examined by us- ing otolith microstructure. Because otolith diameter was correlated to fish length (r=0.73, n=75), we used otolith measures as a proxy for growth rate. Mea- surements were taken inward from the edge of the otolith to the seventh ring. This distance corre- sponded to growth during our seven days of food supplementation. Otolith distances (/i) were then converted into daily growth rates (mm SL/day), by using the following equation generated from a re- gression of otolith diameter on fish length; Growth ^{iotnlith (//.stance + 0.002) / 0.014J/7. Petrik et al,: Recruitment of Micropogonias undulatus 957 We used a blocked two-factor analysis of variance to test the hypothesis that gi-owth rates did not vary among habitat or food supplementation treatments. Effects of predation on recruitment of croaker to varying habitats The null hypothesis that predation has no effect on recruitment of croaker in sand or seagrass habitats was tested by using cages to limit predator access to experimental plots. On 18 March 1996, a fortuitous seasonal low tide completely exposed the study site to air. allowing us to erect cages and to ensure that no recruits or predators occupied the cages at the start of the experiment. Cages (2 x 2 x 1 m) were constructed with 25-mm mesh on four sides, whereas cage controls had mesh on two sides. Mesh was large enough to be transparent to croaker recruits, but fine enough to prevent predators from entering the cages. A randomized block design was employed, with habi- tat (sand or grass) and predator access (cage or cage control) as fixed effects. Replicates were placed 8 m apart, and blocks were separated by 10 m. The experi- ment was terminated after 7 d. Drop samplers were used to quantify recruits as described previously. Results Habitat preference and use by newly recruited croaker The density of newly settled croaker (mean TL=14.48, SE=0.15) did not differ among sand, seagrass, or marsh edge habitats (F^ ^=0.86, «=0.44, /3=0.13) (Fig. 1). The density of croaker recruits also did not vary between sites (Fj jy=0.09, «=0.77) and the in- teraction between habitat and site was not signifi- cant (^2,12=1-07, «=0.37). Our observation on the behavior of croaker recruits in mesocosms did not reveal a preference between sand and seagrass habitats (one way /-test, ^=1.64, df=ll.a=0.13). An average of36'::HSE=16> of the time was spent in seagrass, and QA^'i (SE = 16) in sand. Effects of food supply on croaker recruitment and growth in varying habitats The abundance of newly settled croaker differed be- tween experimental habitats (Fi.ii=5.98, a=0.03) with greater recruitment in sand ( X=31.6/m-, SE=6.5) than in seagrass habitats (X=13.7/m-, SE=5.9) (Fig. 2). Conversely, we did not detect a difference (Fj ,,=0.13, a=0.73) in the number of croaker in food- supplemented plots ( X =23.2/ m-, SE=8.3 ), compared 12 n •S 4 □ Edge 0 Grass ■ Sand Habitat P = 0 44 Site P = 0 77 Site 1 Site 2 Figure 1 Atlantic croaker densities (mean +1 SE ) sampled from three estuarine habi- tats; marsh edge (edge), seagrass bed (gi-ass). and barren sand (sand). P val- ues from two-way analysis of variance. 40 o 10 n Food Addition 0 Control Habitat P = 0 03 Food P = 0 73 n = 5 Grass Sand Figure 2 Atlantic croaker density (mean +1 SE) in 1-m- artificial seagrass (grass) and sand habitats, with (food addi- tion) and without ( control ) food supple- mentation. P values from a blocked two-factor analysis of variance. with control plots ( X =22. 1/m^ SE=4. 1 ) ( Fig. 2 ). The interaction between habitat and food-supplementa- tion was not significant (Fj jj=0.96, «=0.35 ). The sta- tistical power of this experiment was low ( 1-/3=0.18); 958 Fishery Bulletin 97(4), 1999 0.20 -1 0.18 to ■D ^ 0.16 E 5 2 CD 0.14 0 12 0 1 n Food Addition 0 Control Habitat P = 0 09 Food P = 0 61 Grass Sand Figure 3 Growth rate (mean +1 SEi determined from analysis of otolith microstructure in croaker from l-m- artificial seagrass (grass) and sand habitats with (food addition) and without (con- trol) food supplementation. P values are from a blocked two-factor analysis of variance, and n, the number offish sampled, is given at the base of each bar. however, given the small difference in average croaker density between food-supplementation and control plots (1.1/m-) and the high within-treatment variation, we would have needed 55 replicates to achieve sufficient power (1-/^=0.95) to accept the null hypothesis of no difference between treatment means. Growth rates of newly recruited croaker were the same in sand and grass habitats {F-^ ^0=2. 79, a=0.09, /3<0.001 ), as well as with or without food supplemen- tation (Fi^„=0.26, a=0.61, /j=0.004) (Fig. 3). The in- teraction between habitat and food-supplementation on the growth rate of newly recruited croaker was not significant (Fj 7,1=0.49, a=0.49). Growth averaged 0.148 mm SL/day (SE=0.03) in grass and 0.158 mm (SE=0.03)in sand. Average growth rates of 0.1 53 mm/ day ( SE=0.03) were obsei-ved for both control and food addition treatments (Fig. 3). These growth rates are similar to growth rates reported elsewhere ( Warlen, 1981; Cowan, 1988; Nixon and Jones, 1997), suggest- ing that our back calculation of growth rates from otolith measures were not seriously biased. Effects of predation on recruitment of croaker to varying habitats When we examined recruitment of croaker to experi- mental plots with or without predator access, we were 45 40 - 35 - 30 - 25 20 10 5 - I I Predator Excluded 0 Control Habitat P = 0 34 Predator P = 0 92 n = 5 Grass Sand Figure 4 Atlantic croaker density (mean -t-l SE) in 1 m- artificial seagrass (grass) and sand habitats with predators excluded and allowing preda- tory fish and decapod access (control). P values from a blocked two-factor analysis of variance. unable to detect an effect of predation on croaker recruitment (Fj i2=0.01, a=0.92) (Fig. 4). Croaker density averaged 10.7/m- (SE=3.8) in the caged rep- licates, and 15.8/m- ( SE=10. 1 ) in cage controls ( Fig. 4 ). In contrast with the food supplementation experi- ment, we did not detect a difference in recruitment between grass and sand habitats (Fj j,;=0.96, a=0.34) (Fig. 4). The interaction between habitat and preda- tor access was also not significant (Fj j2=0.10, a=0.76). This experiment also suffered from low power ( 1-/^=0.06 ). Sufficient power to accept the null hypothesis of no difference in croaker density be- tween cage and cage-control treatments (l-/i=0.95), would have required 550 replicates. Discussion Recruitment of fishes with open populations is af- fected by variability in larval supply (Jenkins et al., 1996; Hamer and Jenkins, 1997), habitat selection by settling larvae (Bell et al., 1987) and postsettle- ment mortality (Orth et al., 1984), growth (Levin et al., 1997), and migration (Sogard, 1989). Understand- ing how these processes interact with each other to determine population size has been a major focus of researchers on tropical and temperate reefs (Doherty and Williams, 1988; Caley et al., 1996) and recently in seagrass meadows (Bell et al., 1987; Jenkins et al., 1996; Hamer and Jenkins, 1997). There has also Petrik et al.: Recruitment of Micropogontas undulatus 959 been a growing awareness that understanding how demographic processes vary with habitat structure will be critical for predicting population size in fishes that occur in heterogeneous habitats (Levin, 1994). In this study we examined patterns of abundance of newly settled Atlantic croaker and demonstrated that these fish use different estuarine habitats similarly. The results of our mesocosm experiment suggest that the pattern of similar recruitment in different habi- tats results from a lack of preference for specific habi- tats. In addition, when we investigated habitat dif- ferences in postsettlement growth or survivorship, we were unable to detect strong consequences of us- ing one habitat over another. We found no evidence suggesting that habitat se- lection by settling larvae and habitat-specific postsettlement mortality are important in determin- ing population size in croaker; however, this conclu- sion is based on nonsignificant statistical tests rather than explicit acceptance of the null hypothesis of no difference between treatments. Recent reviews have stressed the importance of power analysis in detect- ing a type-II errors (Peterman, 1990; Reed and Blaustein, 1995; Thomas and Juanes, 1996), and because we wished to draw conclusions from "nega- tive" results, power analysis was particularly impor- tant. In this study, when a difference between treat- ment means was not detected, we examined power in an attempt to determine our ability to accept the null hypothesis. If the power of the test was too low to accept the null hypothesis ( /5<0.05 ), the number of replicates required to achieve this power level was calculated. For example, no difference was detected in mean croaker number between plots from which predators were excluded and control plots, but the power of this experiment was low (1-/3=0.06). Suffi- cient power to be able to accept the null hypothesis, would have required 550 replicates. Our level of rep- lication was inadequate because of the extreme vari- ability in croaker densities among experimental plots — variation likely produced by a combination of stochastic settlement and habitat-specific mortality. The effect of this variability was to weaken the power of our experiments to detect small, but real differ- ences among treatments. Although our experimen- tal design precluded the detection of small treatment effects, the high number of replicates required to detect these small effects suggests that other pro- cesses are likely to be more important in determin- ing variability in abundance. Although many estuarine species select vegetated over unvegetated habitat at settlement (Orth et al., 1984), in some cases, initial patterns of settlement have little to do with habitat selection by individual organisms. Settlement may occur at the first suit- able habitat encountered regardless of specific at- tributes of that habitat (Bell and Westoby, 1986). Additionally, current patterns may exclude delivery of competent larvae to some habitats (Morgan et al., 1996); therefore, even ideal habitats may seldom re- ceive recruits. In such cases, larvae do not select against a habitat, instead that habitat is never an available choice. By experimentally providing habi- tats, and by using a blocked sampling design such that all habitats were available in a particular loca- tion, we eliminated the possibility that settling croaker would not have the opportunity to choose a habitat. In our field sampling and experiments, croaker had the opportunity to choose between veg- etated and unvegetated habitats, but they did not consistently choose one habitat over another. By con- trast, in an identical experiment performed at the same time and in the same study site, pinfish (Lagodon rhomboides) showed strong responses to habitat, food supply, and predators (Levin et al., 1997). Pinfish occurred in much higher densities in vegetated than in unvegetated habitats and also gi'ew faster in grass habitats supplemented with food than in unsupplemented or unvegetated habitats. In ad- dition, the presence of predators reduced pinfish numbers by 50%. The pinfish and croaker occupying experimental plots were similar in size ( 15-25 mm SL), and at this size the diets of the two species are similar (Darcy, 1985; Soto et al., 1998). Thus, it is likely that the lack of response by croaker to the habi- tat attributes we investigated is the result of char- acteristics of the species rather than an artifact of sampling or experimental design. Selection for specific habitats at settlement may overwhelm variation in larval supply, thus produc- ing variability in recruitment that is associated with the preferred habitat. This appears to be the case for pinfish (Levin et al., 1997). Although croaker of- ten form part of fish assemblages within seagi'ass (Rooker et al., 1998), they appear to have broad mi- crohabitat preferences, and our results suggest that there is no strong fitness consequences for croaker using vegetated versus unvegetated habitats. As a result, resources associated with the benthic habi- tat seem unlikely to determine population size of newly recruiting croaker. Rather, where and when larvae that are competent to settle are delivered should determine population size in croaker. The contrasting results for croaker and pinfish may re- flect a more general difference in the processes de- termining population sizes of fish. For fishes, such as pinfish, where settling larvae select specific habi- tats and postsettlement processes reinforce initial settlement patterns, spatial and temporal variabil- ity in habitat should be a strong predictor of future 960 Fishery Bulletin 97(4), 1999 population size. Habitat generalists, such as croaker, can occupy a range of habitats in any particular lo- cation, and thus variability in a specific resource may not determine population size. Consequently, pro- cesses affecting larval supply will be more important predictors of population size than resource-related characteristics. Generalizations about the relative importance of processes affecting recruitment in fishes have been elusive (Caley et al., 1996). Although it is widely rec- ognized that understanding the roles of both pre- and postsettlement is critical, there is still little consen- sus on the relative importance of different processes in determining population size. We suggest that ne- glecting the behavioral and ecological characteris- tics of individual species may be a major obstacle in reaching widely accepted generalizations about pro- cesses affecting recruitment. By performing similar experiments on different species we may uncover generalizations about processes affecting fish popu- lations that have thus far been difficult to attain. Acknowledgments We greatly appreciate the assistance in the field pro- vided by T. P. Good and comments on the manuscript by B. Finley and J. Rooker. Support for this project was provided by Texas SeaGrant NA56RG0388 project R/F-67 with supplemental support from a MARFIN grant from NOAA and NSF grant DEB- 9610353. Literature cited Bell, J. D., and M. Westoby. 1986. Abundance of macrofauna in dense seagrass is due to habitat preference not predation. Oecologia 68:205- 209. Bell. J. D., M. Westoby, and A. S. Steffe. 1987. Fish larvae setthng in seagrass: do they di.scriminate between beds of different leaf density? J. Exp. Mar. Biol. Ecol. 111:133-144. Caley, M. J., M. H. Carr, M. A. Hixon, T. P. Hughes, G. P. Jones, and B. A. Menge. 1996. Recruitment and the local dynamics of open marine populations. Ann. Rev. Ecol. Syst. 27:477-500. Carr, M. H. 1991. Habitat selection and recruitment of an assemblage of temperate zone reef fishes. J. Exp. Mar. Biol. Ecol. 146:113-137. Cowan, J. H., Jr. 1988. Age and growth of Atlantic croaker, Mnriipof^onias iindiilalus, larvae collected in the coastal waters on the northern Gulf of .Mexico as determined by increments in saccular otoliths. Bull. Mar. Sci. 42(31:349-357. Cowan, J. H., and R. F. Shaw. 1988. The distribution, abundance, and transport of larval sciaenids collected during winter and early spring from the continental shelf waters off west Louisiana, Fish. Bull. 86(1):129-142. Darey, G. H. 1985. Synopsis of biological data on the pinfish, Lagodon rhomhoides lPi,sces: Sparidaei. U.S. Dep. Commer., NOAA Technical Report NMFS 23, FAO Fisheries Synopsis 141. Doherty, P. J., and D. McB. Williams. 1988. The replenishment of coral reef fish populations. Oceanogr. Mar. Biol. Ann. Rev. 26:487-551. Fonseca, M. S., W. J. Kenworthy, D. R. Colby, K. A. Rittmaster, and G. W. Thayer. 1990. Comparisons of fauna among natural and trans- planted eelgrass Zostera marina meadows: criteria for mitigation. Mar. Ecol. Prog. Ser. 65:251-264. Forrester, G. E. 1990. Factors influencing the juvenile demography of a coral reef fish. Ecology 71:1666-1681. Hamer, P. A., and G. P. Jenkins. 1997. Larval supply and short-term recruitment of a tem- perate zone demersal fish, the King George whiting, Stllaginodes punctata Cuvier and Valenciennes, to an embayment in south-eastern Australia. J. Exp. Mar. Biol. Ecol. 208(1-2):197-214. Heck, K. L., Jr., and R. J. Orth. 1980. Structured components of eelgrass iZostera marina i meadows in the lower Chesapeake Bay: decapod crusta- ceans. Estuaries 3:289-295. Heck, K. L., and T. A. Thoman. 1981. Experiments on predator-prey interactions in veg- etated habitats. J. Exp. Mar. Biol. Ecol. 53:125-134. Hixon, M. A., and J. P. Beets. 1993. Predation, prey refuges and the structure of coral- reef fish assemblages. Ecol. Monogr. 63( 1 ):77-101. Jenkins, G. P., M. J. Wheatley, and A. G. B. Poore. 1996. Spatial variation in recruitment, growth, and feed- ing of postsettlement King George whiting, Sillaginodes punctata, associated with seagrass beds of Port Phillip Bay, Australia. Can. J. Fish. Aquat. Sci. 53:350-359. Johnson, G. D. 1978. Micropogonias undulatuf: I Linnaeus), Atlantic croaker. U.S. Fish and Wildlife Semce FWS/OBS-78/12, p. 227-233. Jones, G. P. 1997. Relationships between recruitment and postrecruit- ment processes in lagoonal populations of two coral reef fishes. .J. Exp. Mar. Biol. Ecol. 213:231-246. Lassuy, D. R. 1983. Atlantic croaker. Species profiles: life histories and environmental requirements (Gulf of Mexico). U.S. Fish and Wildlife Service FWS/OBS-82-11.3. Levin, P. S. 1991. Effects of microhabitat on recruitment variation in a Gulfof Maine reef fish. Mar Ecol. Prog. Ser. 75(2-31:183- 189. 1994. Small-scale recruitment variation in a temperate fish: the roles of macrophytes and food supply. Environ. Biol. Fish. 40:271-281. Levin, P., R. Petrik, and J. Malone. 1997. Interactive effects of habitat selection, food supply and predation on recruitment of an estuarine fish. Oecologia 112:5.5-63. Morgan, S. G., R. K. Zimmer-Faust, K. L. Heck Jr., and L. D. Coen. 1996. Population regulation of blues crabs Calltnectes aapulux in the northern Gulf of Mexico: postlarval supply. Mar Ecol. Prog. Ser. 133:7.3-88. Petrik et al. Recruitment of Miaopogonias undulotus 961 Nelson, W. G. 1979. Experimental studies of selective predation on am- phipods: consequences for amphipod distribution and abundance. J. Exp. Mar. Biol. Ecol. 38:225-245. Nixon, S. W., and C. M. Jones. 1997. Age and growth of larval and juvenile Atlantic croaker. Micropogoniafi undulatus. from the Middle Atlan- tic Bight and estuarine waters of Virginia. Fish. Bull. 95(4):773-784. Norcross, B. L. 1991. Estuarine recruitment mechanisms of larval Atlan- tic croakers. Tran. Am. Fish. Soc. 120(6i:673-683. Orth, R. J., K. L. Heck, and J. vanMonfrans. 1984. Faunal communities in seagrass beds: a review of the influence of plant structure and prey characteristics on predator-prey relationships. Estuaries 7( 4a ): 339-350. Peterman, R. M. 1990. Statistical power analysis can improve fisheries re- search and management. Can. J. Fish. Aquat. Sci. 47:2-15. Pulich, W. M., and W. A. White. 1991. Decline of submerged vegetation in Galveston Bay system: chronology and relationships to physical processes. J. Coast. Res. 7: 1125-1138. Reed. J. M., and A. R. Blaustein. 1995. Assessment of "nondeclining" amphibian populations using power analysis. Cons. Biol. 9(51:1299-1300. Rooker, J. R., S. A. Holt, M. A. Soto, and G. J. Holt 1 998. Postsettlement patterns of habitat use by sciaenid fishes in subtropical seagrass meadows. Estuaries 21:318—327. Shulman, M. J. 1984. Resource limitation and recruitment patterns in a coral reef fish assemblage. J. Exp. Mar. Biol. Ecol. 74:85-109. Sogard, S. M. 1989. Colonization of artificial seagrass by fishes and de- capod crustaceans: importance of proximity to natural seagrass. J. Exp. Mar. Biol. Ecol. 133:15-37. Soto, M. A., G. J. Holt. S. A. Holt, and J. Rooker. 1998. Food habits and dietary overlap of newly settled red drum {Sciacnnps ocellatus) and Atlantic croaker (Micrnpo- gonias undiilatus) from Texas .seagrass meadows. Gulf Research Reports 10:41-55. Stoner, A. W. 1982. The influence of benthic macrophytes on the forag- ing behavior of pinfish Lagodon rhomboides > L. ). J. Exp. Mar. Biol. Ecol. 58:271-284. Thomas, J. L., R. J. Zimmerman and T. J. Minello. 1990. Abundance patterns of juvenile blue crabs (Catli- nectes sapidiis) in nursery habitats of two Texas bays. Bull. Mar. Sci. 46( 1):115-125. Thomas, L., and F. Juanes. 1996. The importance of statistical power analysis: an ex- ample from Animal Behavior. Anim. Behav. 52:856-859. Tolimieri, N. 1995. Effects of microhabitat characteristics on the settle- ment and recruitment of a coral reef fish at two spatial scales. Oecologia 102:52-63. Tupper, M., and R. G. Boutilier. 1995. Effects of habitat on settlement, growth, and post- settlement survival of Atlantic cod (Gadus nior/iua). Can. J. Fish. Aquat. Sci. 52:1834-1841. Warlen, S. M. 1981. Age and growth of larvae and spawning time of At- lantic croaker larvae in North Carolina. Proc. Annu. Conf. SE Assoc. Fish Wildl. Agencies 34:204-214. Wellington, G. M. 1992. Habitat selection and juvenile persistence control the distribution of two closely related Caribbean damselfishes. Oecologia 90:500-508. White, M. L., and M. E. Chittenden Jr. 1977. Age determination, reproduction, and population dynamics of the Atlantic croaker, Micropogonias undulatus. Fish. Bull. 75(11:109-123. Zimmerman, R. J., T. J. Minello, and G. Zamora. 1984. Selection of vegetated habitat by brown shrimp Panaeus aztecus in a Galveston Bay salt marsh. Fish. Bull. 82: 326-336. 962 Abstract.— Nearshore and shelf fish communities were studied in three ar- eas of lower Cook Inlet, Alaska: the Barren Islands (oceanic and well-mixed waters), Kachemak Bay (mixed oceanic waters with significant freshwater run- offi, and Chisik Island (estuarine wa- ters). Fish were sampled with beach seines ^n = 413 sets) and midwater trawls (n=39sets). We found that lower Cook tnlet supported a diverse near- shore fish community of at least 52 spe- cies. Fifty of these species were caught in Kachemak Bay, 24 at Chisik Island, and 12 at the Barren Islands. Pacific sand lance dominated Barren Islands and Kachemak Bay nearshore habitats, comprising gS^f and Tl'/f of total indi- viduals, respectively. The nearshore Chisik Island fish community was not dominated by any one species; instead it exhibited higher diversity. These spa- tial differences appeared linked to lo- cal oceanographic regimes and sedi- ment influx. Analysis of historical data revealed that the nearshore Kachemak Bay fish community changed signifi- cantly between 1976 and 1996, show- ing increased diversity and abundance in several taxa, notably gadids, salmo- nids, pleuronectids, and sculpins. Decadal differences appeared to be re- lated to large-scale climate changes in the North Pacific. Catches of most taxa peaked in May-August, and were low during other months of the year Sev- eral species were present for only part of the summer. Species composition of seine catches differed significantly be- tween consecutive high and low tides, but not between consecutive sets or years. Midwater trawls took 26 species, 14 of which were present in Kachemak Bay, 19 near Chisik Island, and 7 at the Barren Islands. Community structures in shelf and nearshore waters were similar: diversity was high and abun- dance low at Chisik Island, whereas a few abundant species dominated at both Kachemak Bay and the Barren Islands. In addition, the low fish abun- dance near Chisik Island appeared to be related to declining seabird numbers at this colony. Temporal and geographic variation in fish communities of lower Cook Inlet, Alaska Martin D. Robards John F. Piatt Alaska Biological Science Center Biological Resources Division U S Geological Survey 101 1 E Tudor Road Anchorage. Alaska, 99503 E-mail address (for M D Robards) Martin_Robardsausgsgov Arthur B. Kettle Alaska Maritime National Wildlife Refuge US- Fish and Wildlife Service 2355 Kachemak Bay Drive Homer, Alaska, 99603 Alisa A. Abookire Alaska Biological Science Center Biological Resources Division U S Geological Survey 101 1 E Tudor Road Anchorage, Alaska, 99503 Manuscript accepted 24 March 1999. Fish. Bull. 97(41:962-977 (1999), Dramatic changes in seabird anti marine mammal populations in the Gulf of Alaska have been linked to shifts in abundance and composi- tion of forage fish stocks over the past 20 years (Piatt and Anderson, 1996; Anderson et al.M, Coincident with cyclical fluctuations in seawa- ter temperatures, abundance of sev- eral key forage species, including capelin (Mallotus villosus), prickle- backs (mostly Lumpenella longiro- stris), and Pacific sandfish {Tricho- don trichodon) declined precipi- tously in the late 1970s. Meanwhile populations of large predatory fish, including walleye pollock ( Theragra chalcogramma). Pacific cod (Gadus macrocephalus). and several pleuro- nectids increased dramatically. Cor- respondingly, seabird diets shifted from mostly capelin in the 1970s to mostly Pacific sand lance (Ammo- dytes hexapterus) and juvenile pol- lock by the late 1980s (Piatt and Anderson, 1996). Furthermore, a variety of seabirds and marine mammals exhibited signs of food stress during the 1980s and early 1990s (Piatt and Anderson, 1996), Inferences about changes in Gulf of Alaska fish communities have been based on a limited number of site- specific long-term studies (e.g. Anderson et al.' ); however, evidence is mounting that suggests these changes were wide-spread through- out the region ( Francis et al. , 1998 ). The importance of inshore marine habitats as nursery areas for juve- niles of many marine fish species has been well reported (Poxton et al., 1983; Orsi and Landingham, 1985; Bennett, 1989; Blaber et al., 1995; Dalley and Anderson, 1997), Studies in other parts of the world Anderson, P. J., S.A. Payne, and B. A. Johnson. 1994. Multi-species dynamics and changes in community structure in PavlofBav, Alaska 1972-1992. Natl. Mar Fish. Serv., NOAA, Kodiak, Alaska. Unpubl. manuscript, 26 p. Robards et al ; Variation in fish communities of lower Cook Inlet, Alaska 963 have also described seasonal variation in shallow water fish assemblages (e.g. Horn, 1980: Allen, 1982; Nash and Gibson, 1982; Nash, 1988; Bennett, 1989). Little is known, however, about annual, seasonal, and daily variation in fish abundance in Alaska. Under- standing the responses of fish populations at this scale are critical for examining finer-scale processes such as trophic interactions. Responses at these time scales also put interdecadal changes into perspec- tive, and it is changes at the decadal scale that are rapidly becoming the focus of fishery oceanographers (Francis etal, 1998). Our primary objectives were to assess variability in abundance, diversity, and distribution of nearshore fishes in three oceanographically distinct locations within lower Cook Inlet and to compare these data with information collected by Blackburn et al. ( 1980) 20 years ago. Until our study, Blackburn et al.'s 1976 sur\'eys provided the only comprehensive description of nearshore fish communities in the region. Con- current sampling of offshore species with midwater trawls also allowed us to make comparisons between shelf and nearshore fish communities. Methods Study sites Lower Cook Inlet (Fig. 1), in south-central Alaska, is the largest embayment in the northern Gulf of Alaska. The area supports several important seabird colonies and numerous marine mammals, as well as important commercial and recreational fisheries for salmon iOncorhynchus spp. ) and Pacific halibut (Hippoglossus stenolepis). We studied three lower Cook Inlet areas that primarily support colonies of fish-eating common murres ( Uria aalge ) and black- legged kittiwakes (Rissa tj-idactyla): Gull Island, in Kachemak Bay; Chisik Island, on the western side of Cook Inlet; and the Barren Islands, near the en- trance to Cook Inlet. Investigations of nearshore fish communities were initiated around these colonies to provide information that can be used to improve un- derstanding of how forage fish abundance and dis- tribution may influence seabird diets and productiv- ity in these areas. Kachemak Bay (Fig. 1) lies along the southeast- ern shore of Cook Inlet. The bay is 38 km wide at its entrance and 62 km long. The upper 6 km consists of mud flats that are exposed during low tide. Depths are relatively shallow, ranging from about 35 to 90 m, with some deeper areas (100 to 165 m) present off Gull Island along the south-central side of the bay. Water entering the bay originates from the Gulf of Figure 1 Map of lower Cook Inlet showing the three study areas, prevailing currents, bathymetry, and both beach seine and midwater trawl locations. Alaska and is largely oceanic (Fig. 1; Burbank, 1977 ). Chisik Island, on the western side of Cook Inlet ( Fig. 1 ), is located in the mouth of Tuxedni Bay, which receives freshwater from glacier-fed rivers. Water passing outside the island is also estuarine, because it receives significant glacier-fed freshwater input from large rivers at the head of Cook Inlet before circulating down the western side of the inlet (Burbank, 1977; Feely and Massoth, 1982). Near- shore habitats around Chisik Island contain few sandy substrates and consist primarily of glacial silt and mud flats interspersed with rocky substrates that are exposed at low tides. The Barren Islands, near the entrance to Cook Inlet (Fig. 1), are situated in a transition zone be- tween deep Gulf of Alaska waters and the shallow Cook Inlet estuary. The Alaska Coastal Current en- ters Cook Inlet north of the Barren Islands, leading to intense upwelling of cold, nutrient-rich waters onto the shallow southeastern Cook Inlet shelf (Burbank, 1977). Because of the upwelling and strong tidal ac- tion (second highest tidal range in North America), waters in this area are turbulent and well mixed. 964 Fishery Bulletin 97(4), 1999 Field collections We used beach seines to sample nearshore fish com- munities. These nets effectively and nonselectively sample shallow, inshore waters with sandy or smooth bottoms (Cailliet et al., 1986). In Kachemak Bay, samples were obtained during 16 June-26 July 1995 and 8 February-10 December 1996. Weather and sea conditions prevented sampling in January and No- vember 1996. Sampling at Chisik Island and the Barren Islands occurred during 3 July-17 August, and 26 June-8 September 1996, respectively. Our variable-mesh nets were 44 m long. The middle 15.3-m-long, 4-m-deep section was made of 3-mm knotless nylon stretch mesh (sm), and the wings tapered to a depth of 2.3 m were constructed of 13-mm knotted sm nylon. Thirty meters of rope were attached to the ends for deployment. Blackburn et al. (1980) used similar-size seines to sample Kachemak Bay in 1976. Nets were set parallel to shore about 25 m from the beach as described by Cailliet et al. (1986). Samples were collected about every two weeks in Kachemak Bay during May-Sep- tember, and once a month throughout the winter. Samples from Chisik Island and the Barren Islands were collected every two weeks for the duration of their field seasons. We sampled 38 Kachemak Bay sites that were seined by Blackburn et al. (1980) in 1976. At these sites, we made 60 and 245 sets during May-Septem- ber of 1995 and 1996, respectively. In comparison, Blackburn et al. made 131 sets during May-Septem- ber 1976. Raw data from the 1976 study were obtained from J. Blackburn, Alaska Department of Fish and Game (ADF&G), Kodiak, Alaska. Sets were made at eight sites on the west side of Chisik Island. Twenty-three of these sets were made at high tide and seven at low tide (low tide sets were limited by deep mud in the subtidal zone). At the Barren Islands, 40 sets (18 at high tide and 22 at low tide) were made in Amatuli Cove. Beach seining was conducted within one-hour win- dows on either side of high and low tides. To assess tidal influence on catch rates, seven sites in Kachemak Bay were seined regularly on consecutive high and low tides during periods of maximum tidal oscillation (about every 15 days). A single set usu- ally provides good representation of species richness and dominant species rank (Allen et al., 1992). How- ever, to assess variability of beach seine catches, two consecutive sets were made adjacent to each other at the seven sites intensively studied at each tidal stage. As a result, these sites were sampled four times on each day that seining occurred. Fish were sorted and counted by species. Owing to the key trophic role of sand lance in the Gulf of Alaska (Blackburn and Anderson, 1997), these fish were also separated into adult (age groups >1) and juvenile (age group 0) cat- egories on the basis of otolith interpretations. Midwater trawls were used to sample offshore shelf environments by using the ADF&G 22-m stern trawler RV Pandalus in July 1996. Forage fish were located with a Biosonics DT4000 digital echosounder (120 kHz), and significant targets were fished by using a modified herring trawl with a SO-m^ open- ing. Mesh sizes diminished stepwise from about 5 cm in the wings to 1 cm at the codend. The codend was lined with 3-mm mesh and contained a collecting bucket. Trawls were monitored with a Furuno net- sounding system. Tow duration ranged from 20 to 60 minutes depending on fish concentrations. Midwater trawls were not directly comparable to beach seines; trawls selectively sampled the pelagic zone, whereas beach seines unselectively sampled both pelagic and demersal zones. Analyses We calculated four indices to assess broad differences between sites and sampling periods. The Shannon- Wiener index ( H' ) indicates diversity; it increases as both the species number (richness) and equitability of species abundance (evenness) increase (Pielou, 1977). To measure species "richness" in diversity, we used Margalefs index (D; Margalef 1968), and to assess equitability of species abundance, we calcu- lated Pielou's evenness function (J'; Pielou, 1977). Similarity between species lists from different sam- pling periods was tested by using Jaccard's similar- ity coefficient for presence and absence data ( Boesch, 1977). All diversity calculations were based on num- bers of individuals and were made with natural logs (log,,). Species assemblages were compared statisti- cally with the Mann-Whitney rank sum test. Results Interdecadal comparison of beach seine catches in Kachemak Bay: 1976 versus 1995-96 Beach seines were an efficient method of catching nearshore fish. Of the 305 sets made in Kachemak Bay during May-September in 1995 and 1996, only four failed to yield fish.A total of 155,991 fish compris- ing 50 species were caught in these seines (Table 1). Of these species, 35 were primarily represented by ju- venile stages. All species found during winter were also present during summer Sand lance were most numerous; they represented 71% of total individuals. Robards et al : Variation in fish communities of lower Cool< Inlet, Alaska 965 Table 1 Frequency of occurrence and catch-per -unit-of-effort (CPUE) by species cau ght in beach seines during 1976 (May 21- -September 29) and 1995/96 (May 16-September 27). 1976' (131 Sets) 1995-96 (305 Sets) Change Change % Vf in% in Status Common name Scientific name Occurence CPUE Occurrence CPUE Occurrence-' CPUE^ Abundant >50' / of sets in at Dolly varden Salvelinus malma 51.1 6.03 46.6 5,36 = = least one of the Pacific sand lance Ammodytes hexapterus 40.5 248.02 50.8 363.28 = = time periods Great sculpin Myoxocephalus polyacanthocephalus 32.8 0.92 56.7 1.59 = = Common 10-50'y of sets Pink salmon Oncorhynchus gorbuscha 24.4 6.14 39.0 28.16 = = m at least one of Rock sole Pleuronectes bilineatus 15.3 0.33 30.2 1.24 = = the time periods Pacific herring Ctupea pallasi 14.5 35.62 17.4 64.94 = = Pacific cod Gadus macrocephalus 0.8 0.01 27.2 14.49 +++ +++ Whitespotted greenling Hexagrammos stellen 10.7 0.25 17,0 0.68 = = Tubenose poacher Pallasma barbata 5.3 0.07 15,7 0.42 + = Threespine stickleback Gasterosteus aculeatus 13.0 0.27 6.2 0.14 -- = Chinook salmon Oncorhynchus tshawytscha 12.2 2.20 5.6 0.25 -- = Surf smelt Hypomesus pretiosus 11.5 1.36 4.9 0.16 -- = Silverspotted sculpin Blepsias cirrhosus 0.8 0.01 13.8 0.51 +++ + Occasional Snake prickleback Lumpenus sagitta 6.9 0.41 5.6 0.58 = = 1-109; of sets Chum salmon Oncorhynchus keta 3.1 0.06 7.5 2.58 ++ ++ in at least one of Saffron cod Eteginus gracilis 1.5 0.02 8.9 1.71 +++ + the time periods Pacific staghorn sculpin Lcplocottus armatus 6.1 0.09 3.3 0.04 = = Sockeye salmon Oncorhynchus nerka 0.8 0.01 8.5 2.54 +++ ++ Pacific tomcod Microgadus proximus 0.0 <0.01 9.2 2.16 +++ +++ Crescent gunnel Pholis laeta 2.3 0.04 4.9 0.19 + = Starry flounder Platichthys stellatus 3.8 0.05 2.6 0.15 = = Pacific sandfish Trichodon trichodon 0.0 <0.01 6.2 0.15 +++ +++ Kelp greenling Hexagrammos decagrammus 0.0 <0.01 4.6 0.15 V V Buffalo sculpin Enophrys bison 0.8 0.02 3.6 0.08 +++ = Slender eelblenny Lumpenus fabricii 0,0 <0.01 4.3 0.88 +++ +++ Masked greenling Hexagra m m os octogra m m us 3.8 0.15 0.3 0.01 — - Rock greenling Hexagrammos lagocephalus 0.0 <0.01 3.6 0.15 •>4 •>4 Lobefin snailfish Lipans greeni 0.0 <0.01 3.3 <0.01 + + + + + + Lingcod Ophiodon elongatus 1.5 0.02 1.3 0.02 = = Butter sole Pleuronectes isolepts 1.5 0.02 1.3 0.01 = = Walleye pollock Theragra chalcogramma 0.0 0.00 2.6 4.62 + + + + + + Coho salmon Oncorhynchus kisutch 1.5 0..50 0.7 0.01 - - Warty sculpin Myoxocephalus verrucosus 0.0 0,00 2.0 0.04 + + + + + + Sablefish Anoplopoma fimbria 0.0 0.00 1.6 0.17 + + + + + + Longsnout prickleback Lumpenella longirostns 1.5 0.06 0.0 0.00 — Daubed shanny Lumpenus maculatus 0.0 <0.01 1,3 0,02 + + + + + + cntitniui'd Pacific herring (Cliipea pallasi), pink salmon {Oncorhynchus gorbuscha), walleye pollock, and Pa- cific cod accounted for 22% of the remaining individu- als. Great sculpin (Myoxocephalus polyacantho- cephalus), sand lance, and Dolly Varden (Sa/t'e//ni/s malma) were the most frequently caught species during 1995-1996, each occurring in over 40'}f of the samples. 966 Fishery Bulletin 97(4), 1999 Table 1 (continued) 1976' (131 Sets) 1995/96 (305 Sets) Change Change 7c 'J in fJ in Status Common name Scientific name Occurence CPUE Occurrence CPUE Occurrence- CPUE* Rare < 1% of sets in Flathead sole Hippoglossoides elasxodon 0.0 <0.01 1.3 0.02 +++ +++ both time periods Northern rockfish Sehastes polyspinis 0.8 <0.01 0.7 0.01 .5 5 Sawback poacher Sarritor frenatus 0.8 <0.01 0.3 <0.01 5 5 Soft sculpin Psychrolutes sigalutes 0.0 <0.01 1.0 0.02 .5 5 Petrale sole Eopsetta jordani 0.0 <0.01 1.0 0.04 .5 .5 Prowfish Zaprora silcniis 0.0 <0.01 0,7 0.01 .5 5 Padded sculpin Artedius fenestralis 0.0 <0.01 0.7 0.01 .5 5 Pacific halibut Hippoglossus stenolepis 0.0 <0.01 0.7 0.01 .5 5 English sole Pleuronectes vetulus 0.0 <0.01 0.7 0.01 .5 5 Capelin Mallotus villosus 0.0 <0.01 0.3 <0.01 5 S Arctic shanny Stichaeus piinctatiis 0.0 <0.01 0.3 <0.01 .5 5 Yellow Irish lord Hemilepidotus jordani 0.0 <0.01 0.3 <0.01 5 5 Ribbed sculpin Triglops pingeli 0.0 <0.01 0.3 <0.01 5 5 Smooth alligatorfish Anoptagonus inermis 0.0 <0.01 0.3 <0.01 .5 5 Smooth lumpsucker Aptocyclus ventricosus 0.0 <0.01 0.3 <0.01 5 5 ' 1976 data also inc udes: 28""^ sets with un dentified sculpins and 259( sets with unit entified greenlings. Dat a from J. Blackburn lAlaska Dept Fish and Game. Kodiaki. - Change in frequency of occurrence betweer 1976 and 199.5-96 defined by = i<100'ich inge in fi equencyi. + or -1100-200'; change). ++ or -1200- | 300% change), or +++ or — (>300 '» changei. ' Change in CPUE between 1976 and 1995/96 defined by = lvalues within an order of magnitude). + or - ( values chang ed by one order of magni- tude). ++ or -- (val ues changed by two orders of magnitude), and +++ or " (values ch anged by at least th ree orders of magnitude). "* Unidentified greer lings in 1976 were a mix of rock and kelp greenlings. ■^ CPUE and frequency of occurrence were too small to infer change. In 1976, a total of 39,927 fish comprising at least 28 species were collected from 131 seine sets (green- ling, Hexagrammidae, and sculpins, Cottidae, were not always identified to species; Blackburn et al., 1980; Table 1). To compare these data statistically with 1995-96 information, sculpin and greenling were combined into two general categories. By num- ber, sand lance (81%) and herring (12'^() accounted for more than 93 9( of the 1976 catch. Four of the five species that dominated 1976 catches also dominated the 1995-96 catches (sand lance, herring, Dolly Varden, and pink salmon), and sand lance and her- ring were the two most abundant fish caught during both study periods. In 1976, only four lO.OV'c) fish were gadids (3 saffron cod, Eleginus gracilis; and 1 Pacific cod). By 1995-96, gadids (Pacific cod; saffron cod; Pacific tomcod, Microgadiis proximus; and wall- eye pollock) were more numerous on the basis of catch-per-unit-of-effort (CPUE, defined as catch-per- individual-seine) by 1-3 orders of magnitude and had increased in frequency of occurrence by more than 300Vi (Table 1 ). Furthermore, Pacific cod was a domi- nant species in the 1995-96 catches. Similar in- creases in CPUE or frequency of occurrence were apparent for sockeye and chum salmon (O. ncrka. O. keta ), three sculpin species (silverspotted sculpin, Blepsias cirrhosus; buffalo sculpin, Enophrys bison; warty sculpin, Myoxocephalus verrucosus). Pacific sandfish, two prickleback species ( slender eelblenny, Lumpenus fabricii; daubed shanny, L. maculatus), lobefin snailfish (Liparis greeni). sablefish {Anoplo- poma fimbria ). and a flatfish species (Hippoglossoides elassodon ). Marked declines only occurred in catches of masked greenling (Hexagrammos octogrammus), coho salmon (O. kisutch ), and longsnout prickleback (Lunipcnella longirostris). Differences between 1976 and 1995-96 catches were significant on the basis of change in CPUE (Mann-Whitney rank sum test; /=1339.5, P=0.007), percent composition (/=1373.0, P=0.018), and frequency of capture (/= 1398.0, P=0.033). Changes in fish abundance were accompanied by changes in community diversity (Table 2). The Shannon-Wiener index (Hi was greater in 1995-96 than in 1976, refiecting the large increase in gadid species during the 1990s (accounting for two thirds of the difference in H). Species richness (D) was also higher in 1995-96, compared with 1976. This differ- ence resulted from the fact that four species repre- sented 92^/^ of the total 1995-96 catch. In contrast, only two species accounted for 93''^ of the 1976 catch. l! Robards et al : Variation in fish communities of lower Cook Inlet, Alaska 967 Table 2 Catch-per-unit-of-effort (CPUE) and species diversity indices from nearshore and shelf areas of Kachemak Bay (1976 and 1995-96). Chisik Island, and the Barren Islands. Diversity indices: H' = species diversity; J' = species evenness function; D = species richness (see "Methods" section). Location Kachemak Bay Kachemak Bay Chisik Island Barren Islands CPUE H' D Year Nearshore Shelf 1976 1995-96 1996 1996 .305 511 33 4506 345 92 821 Nearshore Shelf Nearshore Shelf Nearshore Shelf 0.74 0.23 — 2.27 — 1.05 1.07 0.29 0.40 3.26 1.51 2.13 1.89 0.67 0.64 3.34 2.85 0.06 0.31 0.03 0.16 0.91 0.63 Table 3 Comparison of 1995 and 1996 Kachemak Bay beach seine catches for species that occurred in more than 10^7^ of sets. Species 1995 Frequency of capture Percent of total catch CPUE 75.0 66.6 310.2 68.3 1.8 8.5 68.3 0.7 3.2 51.7 3.0 14.0 33.3 0.2 1.0 41.7 7.4 34.4 30.0 16.1 75.1 20.0 0.2 0.9 31.7 0.3 1.3 20.0 0.2 0.8 1996 Frequency of capture Percent of total catch CPUE Pacific sand lance Dolly varden Great sculpin Pink salmon Rock sole Pacific cod Pacific herring Whitespotted greenling Tubenose poacher Silverspotted sculpin 67.2 55.7 49.2 52.5 36.9 18.9 11.5 21.3 11.5 17.2 60.0 1.8 0.2 13.8 0.4 0.5 16.5 0.1 >0.1 0.1 250.1 7.5 1.0 57.4 1.5 2.2 68.8 0.5 0.2 0.5 The low equitability (J' ) in both time periods reflected dominance by a few species (e.g. sand lance, herrmg) in catches. Jaccard's similarity coefficient indicated only mod- erate (59*^) similarity in the presence-absence of species between 1995-96 and 1976. This may have resulted from a combination of 1 ) a dramatic increase in abundance of some species in 1995-96 (e.g. tom- cod, sandfish); and 2) increased fishing effort in 1995- 96, which may have increased catches of less com- mon species (e.g. prowfish, Zaprora silenus; capelin) that might have also been detected in 1976, if fish- ing effort had been greater. Interannual comparison of beach seine catches in Kachemak Bay: 1995 versus 1996 As evidenced by catches of common species in Kachemak Bay (Table 3), there was little obvious difference in catches between consecutive years. During June and July of 1995, 27,944 fish belonging to 42 species were caught in 60 beach seine sets ( CPUE=466). During the same period in 1996, 50,859 fish representing 37 species were found in 122 sets (CPUE=417 ). Forty-four species were identified over the course of the two year study, of which 30 were caught in both years. Species diversity ( H' ) was simi- lar in 1995 and 1996 ( 1.21 and 1.22, respectively), as were species richness (D; 3.61 and 3.43, respectively) and equitability (J'; 0.34 in both years). Jaccard's similarity coefficient indicated a 68% similarity between the species assemblages caught in 1995 and 1996. Species not represented in one year or the other were generally uncommon species that were found in only a few seine sets. Differences in percent composition of 1995 and 1996 species assem- blages were not significant (Mann-Whitney rank sum test). 968 Fishery Bulletin 97(4), 1999 Seasonal variation in the nearshore fish community of Kachemak Bay To examine seasonal variation in nearshore fish abundance, we used only data collected in 1996 ( 130,325 fish, 46 species, 283 beach seine sets in Feb- ruary-December). Most species were present in greatest numbers during summer months, appar- ently-moving inshore as water temperatures in- creased above 4°C in May (Fig. 2), and moving off- shore as temperatures declined in October (Fig. 3). Almost no fish were caught or observed during the winter months (December-March; Fig. 3); however, CPUE increased markedly for all species in June and declined dramatically in mid-July. This pattern was also reported by Blackburn et al. ( 1980; Fig. 4). This change largely resulted from changes in nearshore sand lance abundance but was compounded by move- ments of pink salmon and Pacific cod. Large catches of juvenile pink salmon increased the CPUE in June as they migrated along shorelines (Orsi and Landingham, 1985), but salmon CPUE declined rap- idly in July as fish moved offshore (Fig. 3; Blackburn et al. , 1980). Catches of gadids (particularly Pacific cod and pollock) increased markedly in late summer (Fig. 3) and contributed to the large August increase in CPUE. However, most of the August-September increase in CPUE was due to recruitment of first- year sand lance (Fig. 3), which dominated late sum- mer seine catches after adults moved out of nearshore environments in late July (Blackburn et al. , 1980). In October, adult sand lance returned to inshore waters to spawn (Dick and Warnei', 1982; Robards, u E 12 - ^K 10 - J^ 8 - 3 V 6 - IJ \ 4 ' / \^ 2 - /^ v^ JASON D J F 1996 1 1997 Date M A M J Figure 2 Seasonal variation in 1996 sub.surface temperaturc-s (5 m below low tide |0 m|) at Kachemak Bay i 1 ). Chisik Island (2l, and the Barren Islands (3). 1999). CPUE began declining rapidly in October, in conjunction with a decline in diversity from 34 spe- cies in August to only 3 by December. We also observed strong seasonal trends in use of nearshore habitats by other species. Dolly Varden moved into nearshore waters in April, where they remained throughout July. After July they followed salmon into natal freshwater systems to overwinter (Isakson et al., 1971, Orsi and Landingham, 1985). Juvenile rock sole {Pleuronectes bilineatus) and great sculpins were consistently the most abundant fish found in nearshore habitats during February-March. Flatfish (Pleuronectidae) were caught at low rates throughout spring and summer but were not present after October (Fig. 3). Numbers of juvenile great sculpins increased rapidly in spring owing to an in- flux of small juveniles (<20 mm). Catches of these juveniles fell markedly by late June, and small but regular numbers of second-year individuals were caught during summer. Only one capelin was caught in Kachemak Bay (26 July 1995) prior to October 1996, when 1586 first-year fish were captured in three seine sets. Several large capelin schools were also observed at the same time. Capelin were also captured in three of eight seine sets in December. Although herring were present throughout the sum- mer (Fig. 3), almost all of them (999< ) were captured in five June and four August sets at one area (Hali- but Cove ) historically known for its aggregations of herring (Rounsefell, 1930). Species abundance, diversity (H'), and richness (D) in the nearshore Kachemak Bay fish community in- creased steadily from April to June, peaked in July, and declined rapidly in September. Although sand lance dominated the community in summer, more than 30 species of fish were present during June- August. Species evenness (J') declined throughout summer, when sand lance dominated nearshore habi- tats, but increased again in fall as numbers of spe- cies and individuals declined. Variability among consecutive beach seine sets and tidal states in Kachemak Bay To assess variability in catches among sets and tidal states within Kachemak Bay, we made two sets im- mediately adjacent to each other at each site during consecutive high and low tides (i.e. 4 sets per site, 17 samples, 68 sets). CPUE declined markedly between first and second sets at both high (42''/f decline) and low (50'^ decline) tides, although numbers of species caught remained similar (Table 4). Jaccard's simi- larity coefficient for species composition indicated a high degree of similarity between the two adjacent high (75*^ ) and low (74^7^ ) tide sets. However, species Robards et a\ : Variation in fish communities of lower Cook Inlet, Alaska 969 100 75 diversity, richness, and evenness all increased on the second set, at both tidal states, because of a decrease in dominant species (Table 4). Be- cause species composition of seine catches did not differ significantly between consecutive sets at high and low tidal states (Mann-Whitney rank sum test), it appears that one set is adequate for assessing species richness and dominance, as sug- gested by Allen et al. (1992) Twenty-eight species were repre- sented in high tide sets and the overall CPUE was 425, in contrast to 38 species and a CPUE of only 191 at low tide. This difference prob- ably resulted from a scarcity of pricklebacks (Stichaeidae), gunnels (Pholidae), and sculpins at high tide. The species difference in catches in part accounted for the moderate coefficient of similarity (Jaccard's) between tidal states (58% and 61% on sets 1 and 2, re- spectively). Catch composition dif- fered significantly between high and low tides (Mann-Whitney rank sum test; T=2645.P=0.02). Several schooling species (e.g. pink salmon, adult sand lance) and one demer- sal species (great sculpin) showed little disparity in overall CPUE or frequency of occurrence between tidal states (Table 5). These species (or age classes) appeared to remain close to the shore throughout the tidal cycle. Other species (e.g. Dolly Varden, juvenile sand lance) ap- parently moved from deeper waters into the inter- tidal zone at high tide, as shown by the greater fre- quency of capture and CPUE. Although Pacific cod, saffron cod, whitespotted greenling {Hexagraniiuofi stelleri ), sil verspotted sculpin, juvenile great sculpin, and rock sole were caught at both tide levels, these species appeared to remain preferentially in the subtidal zone during high tide (Table 5). Overall, species diversity, richness, and evenness were great- est at low tides (Table 4). Geographic comparison of beach seine catches from Kachemak Bay, Chisik Island, and the Barren Islands A total of 988 fish representing 24 species were caught in 30 beach seine sets in the warmer (Fig. 2 1 nearshore waters around Chisik Island during sum- Herring «=1 5,305 Salmonidae n= 10.278 Sand lance ^^— Juvenile M=87,610 =^ Adult «=5,068 Pleuronectidae M A M J .1. ASOND FMAMJ Month S O N D Figure 3 Sea.sonal variation in CPUE (bars) and frequency of occurrence (lines) of se- lected fish species caught by beach seines in Kachemak Bay. mer 1996 (Table 6). Dolly Varden was the most com- mon species; it was present in 63% of the sets and represented 30% of total catch by numbers. In con- trast to Kachemak Bay and the Barren Islands, sand lance were found in only 33% of sets and accounted for only 24% of the catch by number. Snake pricklebacks (Lumpenus sagitta; 12% ) and Pacific cod (8% ) were the next most abundant species. Sculpins and flatfishes were also commonly caught in the sets. A total of 180.232 fish representing at least 12 spe- cies (including 482 unidentified sculpins, 1 uniden- tified flatfish, and 1 unidentified greenling) were caught in 40 seine sets in the somewhat cooler (Fig. 2) nearshore waters at the Barren Islands during sum- mer 1996 (Table 6). Barren Islands catches were dominated by sand lance (predominantly juveniles; Table 6). This species was found in 90% of the sets and represented over 99% of the total catch. Other species commonly found in the sets included Pacific 970 Fishery Bulletin 97(4), 1999 cod (>50%), sculpins (>40%), and butter sole (Pleiiro- nectes isolepis; 18%). CPUE varied by about two orders of magnitude among the three areas. On the basis of this variable, we determined that Barren Islands waters were more productive than those of Kachemak Bay and Chisik Island (Table 2). However, Chisik Island and Kachemak Bay displayed greater species diversity and richness than ^e Barren Islands (Table 2; Fig. 5). Equity (J') was much greater at Chisik Island (Table 2), indi- cating that this community was not dominated by any one species. Jaccard's similarity coefficients were low when compared between areas. Species assem- blages in Kachemak Bay and at Chisik Island were similar (42% ), and both of these areas differed from the Barren Islands (<25% similarity). Comparison of nearshore and shelf communities in lower Cook Inlet 1250 ^ 4 4 Number of sets 9 16 26 40 40 14 24 69 53 80 19 Year 1976 1996 M J J A S Month O N D Figure 4 Seasonal variation of CPUE in beach seines during 1976 ( ) and 1996 ( ) in Kachemak Bav. In 1996, midwater trawls were made during July at all three study sites (Fig. 1 ). Nine species were found Table 4 Catch-pe ■•-unit-of-e Tort 1 CPUE lands pecies diversi tv in- dices for consecutive beach seine sets made at high and low tides in Kachemak Bay. D versity indices H' = species diversity D = species richness ; J' = species evenness func- tion (see 'Methods" section). Tidal state Set CPUE Species H' D J' High 1 539.35 24 0.85 2.34 0.27 High 2 310.97 25 1.42 2.59 0.44 High Overall 425.16 28 1.09 2.63 0.33 Low 1 255.26 33 1.96 3.53 0..56 Low 2 127.71 33 2.29 3.82 0.65 Low Overall 191.49 38 2.17 3.91 0.60 Table S Frequency of occurrence and catch-per-unit-of-effort (CPUE) of selected fish s pecies (occurring in more than lOVc of sets) caught on consecutive beach seine sets made at high and 1 3w tide in Kachemak Bay. Species High tide Low tide Set 1 (34 sets) Set 2 (34 sets) Set 1 (34 sets) Set 2 (34 sets) % Occurrence CPUE % Occurrence CPUE % Occurrence CPUE 'i Occurrence CPUE Pink salmon 47.1 77.6 38.2 44.5 44.1 41.68 38.2 24.24 Dolly varden 47.1 3.09 47.1 14.85 29.4 1.65 32.4 2.26 Pacific cod 8.8 3.24 11.8 6.32 29.4 22.82 26.5 18.85 Saffron cod 5.9 0.06 2.9 0.74 14.7 4.88 14.7 7.82 Juvenile sand lance 44.1 417.94 61.8 187.32 32.4 88.97 32.4 9.79 Adult sand lance 35.3 8.74 32.4 16.4 41.2 24.76 38.2 2.00 Whitespotted greenling 8.8 0.15 2.9 0.03 20.6 0.32 23.5 2.06 Silverspotted sculpin 0.0 0.00 5.9 0.12 26.5 2.06 29.4 0.59 Juvenile great sculpin 8.8 5.88 5.9 15.24 26.5 46.59 23.5 33.74 Adult great sculpin 41.2 0.94 52.9 1.53 58.82 I 91 44,12 0.85 TubenosG poacher 2.9 0.03 2.9 0.09 8.8 0.18 11.8 0.32 Rock sole 2.9 0.03 11.8 0.12 58.8 3.32 .50.0 2.59 Robards el al : Variation in fish communities of lower Cook Inlet, Alaska 971 (Table 7) that were not present in any of the beach seine sets (Tables 1 and 6). Similar proportions of the most abundant taxa were evident in trawl and beach seine samples from Kachemak Bay (Fig. 5), and sand lance dominated both nearshore and off- shore catches. Few midwater trawls were made at Chisik Island because acoustic sign signals rarely indicated that fish were present in the water column. There was moderate overlap in composition of seine and trawl catches at Chisik Island, and no single species dominated the samples (Fig. 5). In contrast, the shelf environment of the Barren Islands was clearly dominated by walleye pollock, and nearshore habitats were almost completely populated by sand lance. CPUE (defined as catch-per-individual-trawl) for trawl sets varied by an order of magnitude among the three study areas, paralleling results from beach seine sets (Table 2). Geographic trends in commu- nity diversity indices from trawl catches were also similar to those calculated for nearshore beach seine sets (Table 2). Species diversity (H'), species richness (D), and equity (J') were highest at Chisik Island, somewhat lower in Kachemak Bay, and lowest at the Barren Islands (Fig. 5). Table 6 Frequency of occurrence and catch-per- unit-of-effort (CPUE) by species for beach seine catche 5 at Chisik Island (30 sets and the Barren Islands (40 sets) in 1996, Common name' Chisik Island (988 fish) Barren Islands (180.232 fish) % Occurrence CPUE % Occurrence CPUE Pacific sand lance 33.3 7.77 90.0 4465.03 Pacific cod 16.7 2.47 52.5 12.58 Dolly varden 63.3 9.90 2.5 0.13 Great sculpin 50.0 1.07 0.0 0.00 Pink salmon 13.3 0.70 35.0 11.23 Starry flounder 46.7 1.70 0.0 0.00 Unidentified sculpins 0.0 0.00 42.5 12.05 Snake prickleback 33.3 3.97 0.0 0.00 Capelin 13.3 0.43 10.0 3.38 Threespine stickleback •23.3 0.33 0.0 0.00 Butter sole 0.0 0.00 17.5 0.33 Rock sole 16.7 2.07 0.0 0.00 Lingcod 0.0 0.00 15.0 0.33 Whitespotted greenling 13.3 0.73 0.0 0.00 Crescent gunnel 10.0 0.37 0.0 0.00 Kelp greenling 0.0 0.00 7.5 0.13 Pacific herring 6.7 0.60 0.0 0.00 Eulachon 6.7 0.10 0.0 0.00 Silverspotted sculpin 6.7 0.07 0.0 0.00 Padded sculpin 6.7 0.07 0.0 0.00 Pacific staghorn sculpin 6.7 0,07 0.0 0.00 Sawback poacher 6.7 0.17 0.0 0.00 Rock greenling 3.3 0.03 2.5 0.03 Surf smelt 0.0 0.00 5.0 0.55 Coho salmon 3.3 0.03 0.0 0.00 Sockeye salmon 3.3 0.03 0.0 0.00 Longfin smelt 3.3 0.17 0.0 0.00 Pacific tomcod 3.3 0.07 0.0 0.00 Pacific halibut 3.3 0.03 0.0 0.00 Walleye pollock 0.0 0.00 2.5 0.03 ' Latin names included in Table 1 except eL lachon (Thaleichthys pacificus) and longfin smelt iSpirinchus thaleichthys). 972 Fishery Bulletin 97(4), 1999 Table 7 Total numbers of fish by species caught in midwater trawlt in shelf waters within about 40 km of Gull Is and (Kachemak Bay), Chisik Island, and the Barren Islands in 1996. Kachemak Bay Chisik Island Barren Islands Common name Scientific name 16 Sets 6 Sets 17 Sets Walleye pollock Theragra chalcogramma 456 123 12,912 Pacific sand lance Ammodytes hexapterus 3857 132 195 Capelin Mallotus villosus 441 141 840 Pink salmon Oncorhynchus gorbuscha 413 44 0 Pacific cod Gadus macrocephalus 317 4 1 Pacific sandfish Trichodon trichodon 0 59 0 Chinook salmon Oncorhynchus tshawytscha 0 19 0 Tadpole sculpin Psychrolutes paradoxus 16 0 0 Snailfish Cyclopteridae 1 4 0 Eulachon Thaleichthys pacificus 0 10 0 Prowfish Zaprora silenus 9 0 1 Rock sole Pleuronectes bilineatus 2 1 6 Flatfish Pleuronectidae 6 0 0 Armorhead sculpin Gymnocanthus galeatus 0 5 0 Sculpin Myoxocephalus spp. 2 0 1 Sculpin Gymnocanthus spp. 2 0 0 Dover sole Microstomus pacificus 0 2 0 Poacher Bathyagonus spp. 2 0 0 Smooth alligatorfish Anoplagonus inermis 0 2 0 Pacific herring Clupea pallasi 0 1 0 Spinyhead sculpin Dasycottus setiger 0 1 0 Northern sculpin Icelinus borealis 0 1 0 Ribbed sculpin Triglops pmgeli 0 1 0 Arrowtooth flounder Atheresthes stomias 0 1 0 Starry flounder Platichthys stellatus 1 0 0 Pacific lamprey Lampetra tridentata 0 1 0 Total fish 5525 552 13,956 Discussion Nearshore fish communities The overall composition of beach seine catches did not differ between 1995 and 1996, and community indices were also similar between years. This gives us confidence that changes in community composi- tion observed between 1976 and 1995-96 reflected decadal-scale variability rather than just annual variability. The increase in beach seine CPUE be- tween 1976 and 1995-96 may have also been real, but we have less confidence in this result because beach seines were deployed dloser to shore in 1976 (10 m) than in 1995-96 (25 m), and this difference might have influenced catch rates. The most dra- matic difference between decades was in catches of gadids (including pollock). Few were caught in the whole of lower Cook Inlet during 1976, when gadids represented only 0.2*^^ of the total catch (85 individu- als in 262 seines; Blackburn et al., 1980). The 1000- fold increase in gadids that we observed in the mid- 1990s parallels a similar increase in abundance of gadids in offshore shrimp trawls in Cook Inlet (Bechtol, 1997) and the Gulf of Alaska (Piatt and Anderson, 1996; Anderson et al. , 1997 ). Similarly, the increase in pleuronectids and salmonids that we ob- served in Kachemak Bay between 1976 and 1995- 96 was paralleled in the Gulf of Alaska in shrimp trawls ( Anderson et al. , 1997 ) and commercial salmon catches (Francis and Hare, 1994), respectively. Interdecadal changes in abundance of these fishes are probably related to large-scale climate changes in the North Pacific, but causal mechanisms are un- clear (Francis et al., 1998). Water temperatures in the northern Gulf of Alaska changed from being Robards et al : Variation in fish communities of lower Cook Inlet, Alaska 973 100 80 60 40 20 0 c I 80 o I" 60 o '^ 40 c v \> Figure 5 Species composition l^i of total catch I in the nearshorc (■) and shelf ( I areas of Kachemak Bav, Chisik Island, and the Barren Islands. colder than average through the 1970s to warmer than average through the 1980s and 1990s (Royer, 1993). Temperatiu-e changes have been linked to El Nino Southern Oscillation (ENSO) events and shifts in the location and intensity of the Aleutian low-pres- sure cell (Niebauer, 1983). This climatic "regime shift'" caused a major reorganization of North Pa- cific biota (Francis et al., 1998). Phytoplankton and zooplankton abundance increased in the Gulf of Alaska, and this may have increased availability of food to larval stages of salmon and other groundfish, leading to enhanced recruitment (Francis et al., 1998). The timing of peak production of zooplankton biomass also changed, and this may have influenced recruitment in a suite of fish and invertebrates by altering the survival of larvae ("match-mismatch" hypothesis; Anderson and Piatt, in press). However, not all species appear to have been affected by chang- ing climate. It is notable that the most abundant fish jn Kachemak Bay, sand lance, exhibited no signifi- cant change in abundance or frequency of occurrence between 1976 and 1995-96. Sand lance have a unique life history. They spawn during the fall and larvae emerge in January or February, much earlier than the spring plankton blooms or larvae of many other Gulf of Alaska fish species. This strategy may be adaptive in situations where prey availability is un- predictable (Haldorson et al. , 1993). Therefore, sand lance recruitment may have been little affected by changes in the timing or magnitude of zooplankton abundance during spring and early summer. Capelin abundance declined sharply throughout the Gulf of Alaska after the late 1970s (Bechtol, 1997; Piatt and Anderson, 1996) but may have started to rebound in Cook Inlet waters during the early 1990s (Roseneau et al.'-). Apart from catches in 10 beach seine sets ( three sets in Kachemak Bay and at Chisik Island, and four at the Barren Islands), there is little evidence that capelin occur in nearshore areas of Cook Inlet. However, they were the third most com- mon species caught in midwater trawls, and capelin were widely consumed by halibut and seabirds in Cook Inlet in 1995 and 1996 (Roseneau and Byrd, 1997; Roseneau et al.^). This species may occupy habitats offshore of the littoral zone, but largely in- shore of where we made midwater trawls and may therefore be under-represented by this study. The diversity of species in Kachemak Bay is com- parable to that of other subarctic areas of Alaska. Isakson et al. (1971) caught 40 species in the nearshore waters of Amchitka Island and Orsi and Landingham (1985) found 42 species in southeast- ern Alaska. About 20-30 species are typically present in nearshore habitats in the more temperate regions of Sweden (Thorman, 1986a), Norway (Nash, 1988), Scotland (Gibson et al., 1996), and California (Allen and Horn. 1975; Horn, 1980; Allen, 1982). In Kachemak Bay, four species accounted for over 92% of the catch in 1995-96, and two species repre- sented over 937f of the 1976 catch. One species (sand lance) made up more than 99*7^^ of Barren Islands catch, and even at Chisik Island, where species eq- uity was high, only five species accounted for 79^^ of the total numbers. As might be expected from these patterns of relative abundance, these fish were pre- dominantly juveniles (Gibson et al.,1996) and typi- cally low in the trophic web (Allen, 1982). In estua- - Roseneau, D. R. A. B. Kettle, and G. V. Byrd. 1995 and 1996. Common murrc restoration monitoring in the Barren Islands, Alaska. 1993 and 1994. Ex.xon Valdez Oil Spill Restoration Projects 93049 and 94039. Unpubl. final reports. ' Roseneau, D. G., A. B. Kettle, and G. V. Byrd. 1996 and 1997. Barren Islands seabird studies. In D. C. Duffy (com- piler I, APEX: Alaska Predator Ecosystem E.\periment. Exxon Valdez Oil Spill Restoration Project. Alaska Natural Heritage Pro- gram, University of Alaska, Anchorage. Appendix J. Unpubl. an- nual reports. 974 Fishery Bulletin 97(4), 1999 rine, inshore, and bay habitats in the northeastern Pacific, it appears to be normal for five or fewer spe- cies to account for more than 751 of the individuals in local fish communities, even though the total num- ber of species comprising these communities may be much larger (e.g. Allen and Horn, 1975; Hancock, 1975; Horn, 1980; Allen, 1982; Gordon and Leavings, 1984; Orsi and Landingham, 1985). Although no significant differences were observed in species composition between consecutive beach seine sets at the same study sites, CPUE was mark- edly reduced for the second sets made at both high and low tides. This reduction in CPUE may have resulted from avoidance by schooling species to dis- turbance, or removal of most individuals during the first set. As Gibson et al. (1996) observed, we found marked differences in species composition between tidal states, and the number of species found at low tide was reduced by 26% in sets made at high tide. The fact that only species that typically migrate into the intertidal zone (e.g. pink salmon, sand lance, great sculpin) were caught at high tide may have accounted for this difference. Juvenile sand lance occurred in greater numbers and were caught at higher frequencies during high and flood tides, com- pared with low and ebb states as observed by Blackburn and Anderson (1997). These results sug- gested that juvenile sand lance may have moved lat- erally on each tide, by ascending from demersal habi- tats to pelagic zones at flood stages, and returning at ebb stages. Larval plaice (Pleuronectes platessa ) of the North Sea exhibited this kind of behavior in coastal nursery areas (Rijnsdorp et al., 1985). High-latitude fish assemblages, particularly those found in shallow water habitats, are subject to large seasonal variations in temperature and day length. These physical factors impart a strong natural sea- sonality to community structure (Nash, 1988) as ob- served in the nearshore waters of Kachemak Bay. Some fish species move from shallow water habitats to deeper waters in winter when thermal tolerances are exceeded (Allen and Horn, 1975; Allen, 1982; Bennett, 1989). Decreases in catch size between spring and fall peaks have also been observed by many investigators (e.g. Livingston, 1976; Horn, 1980; Allen, 1982; Thorman, 1986b; Methven and Bajdik, 1994). However, the midsummer CPUE de- clines in Kachemak Bay were not accompanied by declines in species diversity, as obser\^ed by Thorman ( 1986b). Reduction in numbers of adult Atlantic sand lance (Ammodytes marinus) has also been noted in midsummer seabird diets (Monaghan et al. , 1996). This change corresponds to the time when predation by chick-rearing seabirds is at maximum; adult sand lance may be responding to the increased presence of predators by avoiding nearshore and surface habi- tats, or they may be coincidentally estivating in preparation for spawning (Sekiguchi et al. , 1976). Geographic variability The relative abundance and distribution offish spe- cies in lower Cook Inlet appears to be largely deter- mined by oceanography and sediment influx. Intense insular upwelling of nutrient-rich waters around the Barren Islands leads to high productivity along fron- tal zones that in turn results in high abundance of fish. Upwelled water entering Kachemak Bay is nu- trient-rich and becomes locally stratified, resulting in the highest primary production recorded in lower Cook Inlet ( Larrance et al. , 1977 ). In Kachemak Bay, high habitat diversity and the plankton prey-base provide the marine environments needed to support an abundant, diverse fish community. As water cir- culates around Cook Inlet to Chisik Island (Fig. 1), sediment loads increase from freshwater glacial run- off (Feely and Massoth, 1982), leading to maximum primary production that is only about one-tenth of that found in Kachemak Bay (Larrance et al., 1977). The lower abundance of sand lance at Chisik Island probably results from a combination of low food avail- ability and deposition of glacial silt and mud (Feely and Massoth, 1982) that blanket most of the local sub- strates. Sand lance are known to require clean, sandy nearshore substrates (e.g. Pinto et al. , 1984), and they appear to favor these habitats; few were caught in the offshore, oceanic waters surrounding the Barren Is- lands. On the north side of the Alaska Peninsula, sand lance also dominated nearshore catches and were caught less frequently offshore (Houghton, 1987). The storm-prone Barren Islands displayed lower species diversity than the other lower Cook Inlet ar- eas. Species inhabiting this more exposed oceanic region may avoid shallow habitats and favor slightly deeper, less disturbed waters (Thorman, 1986a). lead- ing to reduced nearshore diversity (Horn, 1980). Di- versity of fish in the nearshore Kachemak Bay and Chisik Island environments was apparently related to increasing habitat diversity and reduced salinity, as seen elsewhere (Thorman, 1986b; Blaber et al., 1995). The most notable difference observed between nearshore and offshore shelf environments in lower Cook Inlet was that shelf areas were dominated by walleye pollock. Pollock were found in 889( of midwater trawls, compared with only 39^ in beach seine sets. This species does not appear to use nearshore habitats as nursery areas, as does a re- lated species, the Pacific cod (Houghton, 1987). The presence of some pollock in nearshore areas of Kachemak Bay may be related in part to the pres- r I Robards et al : Variation in fish communities of lower Cook Inlet, Alaska 975 ence of oceanic water within the bay. Pollock also frequent Amchitka Island nearshore environments, which are similarly bathed by mixed, oceanic waters (Isakson et al., 1971). Unequivocal declines in seabird populations (pre- dominantly murres and black-legged kittiwakes) at Chisik Island (Slater et al., 1994) may be related to declines of locally abundant forage fish, particularly sand lance. It is possible that historically larger num- bers of capelin (Piatt and Anderson, 1996) or her- ring (or both) (Rounsefell, 1930; Reid, 1971) may have inhabited this region when stocks of pelagic seabirds were higher prior to the mid-1970s. Colder than av- erage temperatures prior to the late 1970s would have favored both of these fish species (Ware, 1995; Frank etal. 1996). Refuge), Minerals Management Service, University of Alaska, Fairbanks (Institute of Marine Science), and the Alaska Department of Fish and Game. All fish were collected under Alaska Department of Fish and Game collection permits. Special thanks are given to J. Blackburn (Alaska Department of Fish and Game) for providing us with raw data from the 1976 Cook Inlet survey, to Stephanie Zuniga for pro- viding beach seine assistance at the Barren Islands, and Nancy Tileston for invaluable library assistance. We also thank D. Black, P. Desjardins, J. Figurski, A. Harding, B. Keitt, K. Mangel, M. Schultz, T. Van Pelt, and S. Zador for their dedicated help with fieldwork and logistics. The manuscript benefited from careful reviews by Paul Anderson and one anonymous reviewer. Limitations of the study Sampling nearshore habitats with beach seines was limited to sandy and cobble substrates. Strong cui'- rents or inshore swells over 0.5 m also prevented ef- fective retrieval of nets. Therefore, fish inhabiting muddy or rocky substrates, mussel (Mytilus ediiUs) and kelp beds, or the surf zone are under-represented in our study. The surf zone is preferentially used by some species because of low numbers of predators and food-rich waters (Bennett, 1989). Burrowing fish, such as sand lance, may also be under-represented in the beach seine data because of their ability to escape under the net (Dick and Warner, 1982; Gordon and Leavings, 1984; Allen et al., 1992; Robards, personal obs). However, sand lance were the most abundant fish caught in Kachemak Bay and the Barren Islands, indicating that beach seines were clearly effective for sampling this species. 'Juvenile cod migrate from deep water habitats during the day to shallower, nearshore waters at night (Methven and Bajdik, 1994; Gibson et al., 1996). Therefore, cod and other gadids may be un- der-represented in our catches, which were made only during daylight hours. However, diel variability in gadid catches has been shown to be lower than vari- ability associated with tide cycles (Gibson et al.,1996). Acknowledgments Major financial and logistic support for the Cook In- let Seabird Forage Fish Study (CISeaFFS) was pro- vided by the Exxon Valdez Oil Spill (EVOS) Trustee Council (APEX project 97 163m), U.S. Geological Sur- vey (Alaska Biological Science Center), U.S. Fish and Wildlife Service (Alaska Maritime National Wildlife Literature cited Allen, D. M., S. K. Service, and M. V. Ogburn-Matthews. 1992. Factors influencing the collection efficiency of estua- rine fishes. Trans. Am. Fish. Soc. 121:234-244. Allen, L. 1982. Seasonal abundance, composition, and productivity of the littoral fish assemblage in upper Newport Bay, California. Fish. Bull. 80:769-790. Allen, L. G. and M. H. Horn. 1975. Abundance, diversity, and seasonality of fishes in Colorado Lagoon, Alamitos Bay. California. Estuarine Coastal Mar Sci. 3:371-380. Anderson, P. J., J. E. Blackburn, and B. A. Johnson. 1997. Declines of forage species in the Gulf of Alaska, 1972- 95, as an indicator of regime shift. In B. R. Ba.xter (ed.), Proceedings of the symposium on the role of forage fish in the marine ecosystem p. 531-543. Alaska Sea Grant Col- lege Program Ak-SG-97-01. Anderson, P. J., and J. F. Piatt. In press. Community reorganization in the Gulf of Alaska following ocean climate regime shift. Mar Ecol. Prog. Ser Bechtol, W. R. 1997. Changes in forage fish populations of Kachemak Bay, Alaska, during 1976-1995. In B. R. Baxter (ed.), Proceed- ings of the symposium on the role of forage fish in the marine ecosystem, p. 441-455. Alaska Sea Grant College Program Ak-SG-97-01. Bennett, B. A. 1989. The fish community of a moderately exposed beach on the southwestern cape coast of South Africa and an as- sessment of this habitat as a nursery for juvenile fish. Estuarine Coastal Shelf Sci. 28: 293-305. Blaber, S. J. M., D. T. Brewer and J. P. Salini. 1995. Fish communities and the nursery role of the shal- low inshore waters of a tropical bay in the Gulf of Carpentaria, Australia. Estuarine Coastal Shelf Sci. 40: 177-193. Blackburn, J. E. and P. J. Anderson. 1997. Pacific sand lance (Ammodytes hexapterus) growth, seasonal availability, movements, catch variability, and food in the Kodiak-Cook Inlet area of Alaska. In B. R. Baxter ( ed. ), Proceedings of the Symposium on the Role of Forage Fish in the Marine Ecosystem, p. 409-426. Alaska Sea Grant College Program AK-SG-97-01. 976 Fishery Bulletin 97(4), 1999 Blackburn, J. E., K. Anderson, C. I. Hamilton, and S. J. Starr. 1980. Pelagic and demersal fish assessment in the Lower Cook Inlet estuary system. U.S. Department of Com- merce, NOAA, OCSEAP Final Report, Biological Studies 12:259-602. Boesch, D. F. 1977. Application of numerical classification in ecological investigations of water pollution. U.S. Environmental Protection Agency Ecological Research Series EPA-600/3- 77-033, 115 p. Burbahk, D. C. 1977. Circulation studies in Kachemak Bay and lower Cook Inlet. Vol. 3. Environmental studies of Kachemak Bay and lower Cook Inlet (Trasky, L. L., L. B. Flagg, and D. C. Burbank., eds.). Alaska Dept. Fish and Game, Anchor- age, AK, 207 p. Cailliet, G. M., M. S. Love, and A. W. Ebeling. 1986. Fishes: a field and laboratory manual on their struc- ture, identification, and natural history. Wadsworth Pub- lishing Company, Belmont, California, p. 132-134. Dalley, E. L. and J. T. Anderson. 1997. Age-dependent distribution of demersal juvenile At- lantic cod iCladus morhua) in inshore/offshore northeast Newfoundland. Can. J. Fish. Aquat. Sci. 54:168-176. Dick, M. H. and I. M. Warner. 1982. Pacific sand \ance . Ajiimodytes hexapterus Pallas, in the Kodiak Island group, Alaska. Syesis 15:43-50. Feely, R. A., and G. J. Massoth. 1982. Sources, composition, and transport of suspended particulate matter in Lower Cook Inlet and northwestern Shehkof Strait, Alaska. NOAA Technical Report ERL-4 15 PMEL-34, 2« p. Francis, R. C, and S. R. Hare. 1994. Decadal-scale regime shifts in the large marine eco- systems of the Northeast Pacific: a case for historical science. Fish. Oceanogr. 3:279-291. Francis, R. C, S. R. Hare, A. B. Hollowed, and W. S. Wooster. 1998. Effects of interdecadal climate variability on the oce- anic ecosystems of the NE Pacific. Fish. Oceanogr. 7:1— 21. Frank, K. T., J. E. Carscadden, and J. E. Simon. 1996. Recent excursions of capelin iMallotus villosufi) to the Scotian Shelf and Flemish cap during anomalous hy- drographic conditions. Can. J. Fish. Aquat. Sci. 53:1473- 14S6. Gibson, R. N., L. Robb, M. T. Burrows, and A. D. Ansell. 1996. Tidal, diel and longer term changes in the distribu- tion of fishes on a .Scottish sandy beach. Mar. Ecol. Prog. Ser 130:1-17. Gordon, D. K., and C. D. Leavings. 1984. Seasonal changes of inshore fish populations on Stur- geon and Roberts Bank, Eraser River estuary, British Columbia. Can. Tech. Rep. Fi.sh. Aquat. Sci. 1240:1-12. Haldorson, L., M. Pritchett, D. Sterritt, and J. Watts. 1993. Abundance patterns of marine fish larvae during spring in a southeastern Alaskan bay. Fish. Bull. 91:36- 44. Hancock, M. J. 197.'). A survey of the fish fauna in the shallow marine waters of clam lagoon, Adak, Alaska. MS. thesis, Florida Atlantic University, Boca Raton, FL, 49 p. Horn, M. H. 1980. Diel and .seasonal variation in abundance and diver- sity of shallow- water fish populations in Morro Bay, Cali- fornia. Fish. Bull. 78:759-770. Houghton, J. P. 1987. Forage fish use of inshore waters north of the Alaska Peninsula. In Forage fishes of the southeastern Bering Sea, p. 39-47. Conference proceedings U.S.D.O.I., M.M.S., OCS study MMS87-0017, Isakson, J. S., C. A. Simenstad, and R. L. Burgner. 1971. Fish communities and food chains in the Amchitka area. Biosci. 21:666-670. Larrance, J. D., D. A. Tennant, A. J. Chester, and P. A. Ruffio. 1977. Phytoplankton and primary productivity in the north- east Gulf of Alaska and Lower Cook Inlet. In Environ- mental assessment of the Alaskan Continental Shelf, vol. X. Receptors-fish, littoral, benthos, p. 1-136. Outer Con- tinental Shelf Environmental Assessment Program, Boul- der, Colorado. Final report. Livingston, R. J. 1976. Diurnal and seasonal fluctuations or organisms in a north Florida estuary. Estuarine Coastal Mar. Sci. 4:373- 400. Margalef, R. 1968. Perspectives in ecology theory. Univ. Chicago Press, Chicago, IL, 111 p. Methven, D. A., and C. Bajdik. 1994. Temporal variation in size and abundance of juve- nile Atlantic Cod (Gadus morhua) at an inshore site off eastern Newfoundland. Can. J. Fish. Aquat. Sci. 51: 78-90. Monaghan, P., P. J. Wright, M. C. Bailey, J. D. Uttley, P. Walton, and M. D. Burns. 1996. The influence of changes in food abundance on div- ing and surface-feeding seabirds. In W. A. Montevecchi (ed. ), Studies of high-latitude seabirds. 4. Trophic relation- ships and energetics of endotherms in cold ocean systems, p. 10-19. C.W.S, Occ. Pap. 91. Nash, R. D. M. 1988. The effects of disturbance and severe seasonal fluc- tuations in environmental conditions on north temperate shallow-water fish assemblages. Estuarine Coastal Shelf Sci. 26:123-135. Nash, R. D. M., and R. N. Gibson. 1982. Seasonal fluctuations and compositions of two popu- lations of small demersal fishes on the west coast of Scotland. Estuarine Coastal Shelf Sci. 15:485-495. Niebauer, H. J. 1983. Multiyear sea ice variability in the eastern Bering Sea: an update. .J. Geophys. Res. 88:2733-2742. Orsi, J. A. and J. H. Landingham. 1985. Numbers, species, and maturity stages of fish cap- tured with beach seines during the spring 1981 and 1982 in some nearshore marine waters of southeastern Alaska. U.S. Dep. Commer., NOAA Tech. Mem. NMFS F/ NWC-86, 34 p. Piatt, J. F. and P. Anderson. 1996. Response of common murres to the Ex.\on Valdez oil spill and long-term changes in the Gulf of Alaska marine ecosystem. In S. D. Rice, R. B. Spies, D. A. Wolfe, and B. A. Wright (eds.). Exxon Valdez oil spill .symposium pro- ceedings, p. 720-737. Am. Fish. .Soc. Symp. 18. Pielou, E. C. 1977. Mathematical ecology. .John Wiley, New York. NY, 385 p. Pinto, J. M., W. H. Pearson, and J. W. Anderson. 1984. Sediment preferences and oil contamination in the Pacific sand lance Ammodytes hexapterus. Mar. Biol. 83:193-204. Robards et al.: Variation in fish communities of lower Cook Inlet, Alaska 977 Poxton, M. G., A. Eleftheriou, and A. D. Mclntyre. 1983. Food and growth of 0-group flatfish on nursery grounds in the Clyde Sea area. Estuarine Coastal Shelf Sci. 17:319-337. Reid, G. M. 1971. Age, composition, weight, length, and sex of herring, Cli/pea pallasu, used for reduction in Alaska, 1926- 66. U.S. Dep. Commer., NOAA Tech. Rep. NMFS SSRF 634:1-25. Rijnsdorp, A. D., M. Van Stralen, and H. W. Van Der Veer. 1985. Selective tidal transport of North Sea plaice larvae Pleuronectes platessa in coastal nursery areas. Trans. Am. Fish. Soc. 114: 461-470. Robards, M. D., J. F. Piatt, and G. A. Rose. 1999. Maturation, fecundity and mtertidal spawning of Pacific sand lace lAmmodytes hexaptei'iis) in the northern Gulf of Alaska. J. Fish. Biol. 54:1050-1068. Roseneau, D. R. and G. V. Byrd. 1997. Using Pacific halibut to sample the availability of forage fish to seabirds. In B. R. Baxter (ed.). Proceedings of the symposium on the role of forage fish in the marine ecosystem, p. 231-241 Alaska Sea Grant College Program AK-SG-97-01. Rounsefell, G. A. 1930. Contribution to the biology of the Pacific Herring. Clupea palla.iii. and the condition of the fishery in Alaska. Bull. Bur. Fish. Doc. 1080:226-320. Royer, T. C. 1993. High-latitude oceanic variability associated with the 18.6-year nodal tide. J. Geophys. Res. 98:4639-4644. Sekiguchi, H., M. Nagoshi, K. Horiuchi, and N. Nakanishi. 1976. Feeding, fat deposits, and growth of sand eels in Ise Bay, central .Japan. Bull. .Jap. Soc. Sci. Fish. 42:831-835, Slater, L., J. W. NeLson, and J. Ingrum. 1994. Monitoring studies of Lower Cook Inlet seabird colo- nies in 1993 and 1994. U.S. Fish and Wildl. Serv. Rep., AMNWR 94/17. Homer, AK, 43 p. Thorman, S. 1986a. Physical factors affecting the abundance and spe- cies richness of fishes in the shallow waters of the south- ern Bothnian Sea iSwedeni. Estuarine, Coastal Shelf Sci. 22:357-369. 1986b. Seasonal colonization and effects of salinity and temperature on species richness and abundance offish of some brackish and estuarine shallow waters in Sweden. Holarct. Ecol. 9:126-132. Ware, D. M. 1995. The link between ocean climate and herring recruit- ment: proceedings of the seventh pacific coast herring work- shop. Can. Tech. Rep. Fish. Aquat, Sci. number 2060, 133 p. 978 Abstract.— Natural mortality was es- timated for Rhizoprionodon taylori by using seven indirect methods based on relationships between mortality and life history parameters and one direct method, catch curve analysis. Esti- mated values from indirect methods were 0.60 to 1.65. whereas catch curves produced values of 0.56 for females and 0.70 for males. Demographic analysis was undertaken by using standard life history table techniques. Life history tables where each of the seven indirect estimates of natural mortality for R. taylori produced intrinsic rates of natu- ral increase from -1.297 to 0.212 for an unfished population, and only two of the seven produced positive population growth. The implications of these re- sults are discussed in relation to the accuracy of indirect methods for esti- mating natural mortality for R. taylori and other species of shark. The catch curve method was considered the best estimate of natural mortality and gave an intrinsic rate of increase of 0.27 and a doubling tmie of 2.55 years. The re- sults of life history table analysis with the estimate of natural mortality from the catch curve analysis indicated that a R. taylori population could sustain fishing mortality up to 0.18 if applied evenly over all age classes, or 0.67 if age at first capture was two years. The implications of this study are discussed in relation to sustainability of elasmo- branch stocks, particularly short-lived, fast growing, early maturing species. Mortality estimates and demographic analysis for the Australian sharpnose shark, Rhizoprionodon taylori, from northern Australia Colin A. Simpfendorfer Western Australian Marine Research Laboratones PO Box 20 North Beach, Western Australia 6020 Australia Present address: Center for Shark Research Mote Marine Laboratory 1600 Ken Thompson Parkway Sarasota, Florida 34236 E-mail address colins 5 mole org Manuscript accepted 5 October 1998. Fish. Bull. 97: 978-986 ( 1999). In a paper entitled "Problems in the rational exploitation of elasmo- branch populations and some sug- gested solutions," Holden (1974, p. 137) concluded that "elasmobranch stocks offer very limited opportuni- ties for long-term exploitation." Holden's conclusion was based on the fact that elasmobranch life his- tory traits, such as slow growth, long lifespan, late age at maturity, and low fecundity, make popula- tions very susceptible to recruit- ment overfishing. Holden's hypoth- esis has been supported by evidence from a number of shark stocks that have been rapidly overfished. These stocks include the spiny dogfish off Scotland and Norway (Squalus acanthias; Holden, 1968, 1977), the soupfin shark off California (Galeo- rhinus galeus; Ripley, 1946), the basking shark of the Irish Sea iCeto- rhiniis maximus; Parker and Stott, 1965), the porbeagle of the western Atlantic (Lamna nasus; Casey et al., 1978), and the sandbar and dusky sharks of the western North Atlan- tic (Musick et al., 1993). The ma- jority of these examples are long- lived, slow growing, late maturing, temperate-water species. This group of commonly cited examples does not, however, represent the full range of elasmobranch life history traits and as such represents a bi- ased set of data on which to base conclusions about sustainability of elasmobranch stocks. In particular, the examples lack representatives from tropical areas, especially short-lived, rapidly growing, early maturing species. With adequate information on species with a wide range of life histories, a more accu- rate assessment of the ability of elasmobranch stocks to sustain fish- ing pressure should be possible. In recent years the use of demo- graphic analysis (i.e. life history tables) has become popular in the assessment of elasmobranch popu- lations and their ability to be sus- tainably fished (e.g. Hoenig and Gruber, 1990; Cailliet, 1992; CaiUiet et al., 1992; Cortes, 1995; Cortes and Parsons, 1996; Sminkey and Musick, 1996; Au and Smith, 1997; Cortes, 1998). This style of analysis requires only knowledge of the life history traits of a species. Because demographic models are static rep- resentations of populations with a stable age structure, they have limi- tations in respect to density-depen- dent responses (e.g. density-depen- dent natural mortality I and dynamic processes (e.g. fishing and variable reciTjitment) that can be included in dynamic models. The latter types of II Simpfendorfer; Demographic analysis of Rhizoprionodon taylori 979 Table 1 Methods used to estimate moi taUty foi R/u zopnonodoi taylon from life history parameter relationships. GSI = gonadosomatic index; K = von Bertalanffy gro wth parameter; L^, = : von Bertalanffy growth parameter; M = natural mortality; T = average water temperature; /,„„, = maximum age; x„, = age at maturity; Z = total mortality. Method Relationships Hoenig(1983) ln(Z)= 1.46- 1.01 In «,„„,) Pauly (1980) ln(M) = -0.0066 - 0.279 log (L_,) + 0.6543 log (A") + 0.4634 log (T) Gunderson(1980) M= 4.64 GS/- 0.370 Gunderson and Dygert ( 1988) M = 0.03 + 1.68 GSI Jensen (1996) (age) M .v„, Jensen ( 1996) (growth) M= l.Sii: (theoretical) Jensen (1996) (Pauly) M= \.6K models require abundance data (e.g. CPUE time-se- ries data or fishery-independent sui-veys) that are not available for many elasmobranch populations. In cases where abundance data are not available, demographic models may represent the best available technique for the analysis of stock dynamics. The Australian sharpnose shark, Rhizoprionodon taylori. is an ideal example of a small, short-lived, fast growing, early maturing tropical species ( Simpf- endorfer, 1992, 1993). It is endemic in the inner con- tinental shelf waters of northern Australia between the Northwest Shelf and southern Queensland (Last and Stevens, 1994). It is captured throughout much of its range as bycatch in commercial fishing operations, including gillnet fisheries for barramundi, mackerel, and shark, and prawn trawl fisheries (Simpfendorfer and Milward, 1993; Last and Stevens, 1994). This paper reports the results of mortality estimation and demographic analysis for R. taylori. Mortality was estimated in two ways. First, a number of relations between mortality and life history parameters from the literature were used to evaluate the appropri- ateness of these methods for estimating natural mortality in sharks. Second, catch curve analysis was conducted to produce a direct estimate of mortality. The results of the demographic analyses are used to comment on the probable sustainability of short- lived, fast growing, early maturing, tropical elasmo- branch populations. Materials and methods Estimation of mortality Two approaches were taken to estimate mortality in R. taylori. The first was to employ relationships be- tween life history parameters and natural mortality (M) or total mortality (Z) from the literature. Seven relationships were chosen (Table 1) to investigate variability in their results. All seven were based heavily on data from teleost fish, although most in- cluded some data from elasmobranchs. The most widely used relationship in elasmobranch studies is that of Hoenig ( 1983), which uses a linear function to estimate total mortality from maximum age. Al- though this method estimates total mortality, it was assumed to represent natural mortality for/?, taylori because there was little or no fishing for this species in the study area (Simpfendorfer, unpubl. data). The method of Pauly (1980) uses two parameters of the von Bertalanffy growth curve (L^^ and K) and aver- age temperature to estimate natural mortality. Jensen (1996) reanalyzed Pauly's (1980) data and found that natural mortality could be estimated with the same level of accuracy based only on the value of K. In the same paper Jensen (1996) also gave two other relationships for estimating natural mortality based on life history theory, one based on K and the other on age at maturity. The final two relationships selected used maximum female gonadosomatic in- dex (GSI) as an indicator of reproductive effort to estimate natural mortality. The method of Gunderson ( 1980) was based on only 10 species; Gunderson and Dygert (1988) expanded this to 20 species. Data for calculation of M were taken from Simpfendorfer ( 1992; GSI ) and Simpfendorfer ( 1993; von Bertalanffy param- eters, maximum age, and age at maturity). The second approach to the estimation of mortal- ity was catch curve analysis (e.g. Ricker, 1975; Vetter, 1987). Age data for the catch curves were taken from Simpfendorfer ( 1993). Ages were converted to whole years and the natural log of the number of individuals (In N) was plotted against age for males and females 980 Fishery Bulletin 97(4), 1999 separately. Mortality for each sex was calculated as the negative value of the slope of the regression line for points to the right of, and including, the peak In N value. Ricker (1975) used only points to the right of the peak In N value to calculate mortality; how- ever, the small number of age classes in R. taylori meant that including the peak value would increase the reliability of the estimate. To confirm that the peak hi A/^ value should be included in the regression calculations, linear and quadratic functions were fit- ted to the data. If the quadratic function provided a significantly improved fit as judged by an F test, then it was assumed that the inclusion of the peak In N value introduced significant curvature into the data set and so violated the assumption of catch curve calculations that mortality is constant across all age classes. In this situation the peak In N value was not used in the calculation. As with the Hoenig ( 1983 ) method, the estimate of mortality from the catch cui-ves was taken to be natural mortality because there was little or no fishing pressure on the stock from which the age data were collected. Demographic analysis Demographic analysis ofRhizoprionodon taylori was undertaken with standard life history table meth- ods (e.g. Krebs, 1985). The parameters estimated from the life history table were net reproductive rate (/?(,), generation time (G), intrinsic rate of popula- tion increase (r), and population doubling time (^y, '■ Positive values of /• indicate that a population is able to replace itself and thus will not decline, whereas negative values of r indicate that the population is unable to replace itself and will decline. Values of ;• were calculated by solving the Euler equation (Krebs 1985): Lm, 1 x=(i where .V age; maximum age; the proportion of animals surviving to the beginning of a given age class; and age-specific natality. Age and growth data for the demographic analy- sis were taken from Simpfendorfer ( 1993 ). Maximum age was taken as 6 years for males and 7 years for females, and age at maturity'for males and females was taken to be one year. Because Simpfendorfer (1993) was not able to validate age estimates (but did have supporting marginal increment and length- frequency data), sensitivity tests were run to inves- tigate the influence of the uncertainty of age esti- mates (maximum age 10 years, age at maturity 2 years). Reproductive data were taken from Simpfendorfer (1992). Litter size varied significantly with mater- nal length (Fig. 1; /--=0.33, P<0.05). Because obser- vations of litter size were made throughout the year, the litter size for each age class was calculated from the size of females at the midpoint of the age class. Mature females produce a litter each year. The sex ratio of the embryos was not significantly different from 1:1; therefore the litter size for each age class was halved to give the number of female pups per female. Although R. taylori matures at the age of one, the first litter is not produced until the end of the second year. Thus only females that survive to the end of that age class produce young. To accommo- date this in the life history table, age-specific natal- ity was calculated from the number of animals sur- viving to the end of a given age class (i.e. the num- ber at the start of the next age class). Similarly, the age-specific natality of the final age class was set to zero because it was assumed that no animals sur- vived to the end of the last age class. To assess the difference in results between calculating age-specific natality from the number present at the beginning and end of an age class, separate life history tables were constructed and the results compared. A life history table was constructed for each of the values for natural mortality, calculated as described above, to investigate the sensitivity of demographic parameters to different values. To simulate increased mortality of the youngest age class, a number of other authors have doubled the normal value of M for the first age class (e.g. Hoenig and Gruber, 1990; Smith and Abramson, 1990; Cailliet, 1992; Sminkey and Musick, 1996). A life history table was constructed with double the normal value of M for the first R. taylori age class to test the sensitivity of outcomes to this approach. Fishing mortality (F) was incorporated into the survivorship function of the life history table such that total mortality was the sum of M and F. The critical value of F, at which /• equaled zero (i.e. the level of fishing beyond which the population could not replace itself, F,,), was calculated for each life table. For the calculation of F^., fishing mortality was assumed to be equal for each age class. Negative values of F, occurred when the population was de- clining without fishing and indicated that the popu- lation could sustain no fishing. The main source of fishing mortality on R. taylori in northern Australia is gill nets, which do not normally catch animals until they are at least one year of age. To investigate the effect of the age at which F begins (age at first cap- Simpfendorfer: Demographic analysis of Rhizopnonodon taylori 981 Table 2 Life history table (or Rhizopnonodon taylori based on various estimates of female natural mortality (M). /?„ = net reproducdtive rate; G = generation time: r = intrinsic rate of population increase; ty, = population doubling time; F, = level of fishing beyond which the population can not replace itself. Scenario Source of Af A B C D E F G H Hoenig(1983) Pauly (1980) Jensen (1996) (age) Jensen (1996) (growth) Jensen ( 1996) (Pauly ) Gunderson (1980) Gunderson and Dygert ( 1988) Catch curve (female) M Rn 0.60 1.34 1.65 1.52 1.62 1.49 0.70 0.56 1..547 0.200 0.096 0.130 0.103 0.140 1.123 1.758 2.233 1.442 1.301 1.353 1.312 1.366 2.063 2.304 0.212 -0.869 -1.297 -1.119 -1.256 -1.078 0.057 0.271 t. 3.273 -0.798 -0.534 -0.619 -0.552 -0.643 12.102 2.554 0.140 -0.600 -0.910 -0.780 -0.880 -0.750 0.038 0.179 ture, AAFC) the values of r at different levels of F and AAFC were calculated. F after AAFC was assumed to be constant across all ages. Results Mortality The calculation of M from the various life history relations produced estimates ranging from 0.6 to 1.65 (Table 2). All the methods based on von Bertalanffy growth parameters gave results greater than one. Only the Hoenig (1983) and Ciunderson and Dygert (1988) methods yielded estimates less than 1. The length and age frequency data for R. taylori on which the age analysis of Simpfendorfer ( 1993 ) was based are shown in Figure 2. Estimates of M from the catch curves were 0.698 for males and 0.561 for females (Fig. 3). The first age class (0) for both males and females was excluded from the regression analysis to estimate M because it was lower than, and to the left of, the peak In A^ value. Fitting of quadratic functions to the data did not sig- nificantly improve the fit to the data points selected (male: F=3.86, P=0.188; female: F=1.35, P=0.329), confirming that their inclusion did not violate the assumption relating to constant mortality. Demographic analysis Life history tables were constructed to obtain demo- graphic results for the eight different estimates of M for an unfished Rhizoprionodon taylori population (Table 2). Three of the eight (A, G, and H) gave posi- tive values of r: the methods of Hoenig (1983), 10 S 6 Litter size = 0.019TL- 7.919 500 550 600 650 700 750 800 Total length (mm) Figure 1 Litter size of Rhizoprionodon taylori as a function of mater- nal length. Data from Simpfendorfer ( 1993). Gunderson and Dygert (1988) and the catch curve method. The catch curve and Hoenig (1983) meth- ods produced similar demographic results, with r values between 0.2 and 0.3, and with doubling times of 2.5 to 3.5 years (Table 2).These were considerably higher than for the Gunderson and Dygert (1988) method, which gave an r value of 0.057 and a dou- bling time of 12.1 years. The values of r for the re- maining scenarios (B-F) were all much less than zero (-0.869 to -1.297; Table 2), indicating population decrease even with no fishing. The value of M from the catch curve method was used to test the sensitivity of the life history table values to changes in model structure and age para- meters. There was a large difference in age-specific reproductive rate calculated with a proportion of the 982 Fishery Bulletin 97(4), 1999 200 300 400 500 600 700 Total length (mm) 800 25 20 15 10 5 B i 1 II Age class (years) Figure 2 (A) Length and (B) age-frequency distributions of male (open bars) and female (hatched bars) Rhizo- prionodon taylon from the waters of Townsville, Queensland. The arrow indicates size at birth. population surviving to the beginning of an age class (J) rather than to theenci (I). The value of r increased from 0.271 to 0.600, doubling time decreased from 2.544 years to 1.155 years, and the maximum value of F sustainable by the population (F^) increased from 0.179 to 0.600 (Table 3) when the beginning of an age class was used. Doubling the value of M in the first year of life (I vs. K) caused a reduction in the value of r from 0.271 to 0.001, and population dou- bling time increased from 2.544 to 529.2 years (Table 3). The maximum level of fishing mortality that the population could sustain under such a scenario de- creased from 0.179 to 0.001. Doubling first-year M and also calculating age-specific natality from the number surviving at the beginning of age classes (L) produced results identical to the base case (Table 3). Increasing the maximum age to 10 years resulted in a small increase in /; from 0.271 to 0.276, and de- creased population doubling time from 2.544 to 2.516 4 A 3 •\^ • ^v • 2 \^ 1 ^X 0 • 1 1 1 1 1 1 1 ¥01234561 E C —1 B 3 • \n» * ^\ * 2 \. 1 ^ \. • T» 0 1 1 1 1 1 1 1 0 12 3 4 5 6 Age class (years) Figure 3 Catch curves for (A) male and (B) female Rhizo- prionodon taylon derived from data from Simp- fendorfer (1993). Data points for the first age class were not used to calculate regression lines. years (Table 3). Setting age at maturity at two years instead of one resulted in an r of 0.035, and a popu- lation doubling time of 19.82 years (Table 3). Values of r varied considerably with changes in F and AAFC (Fig. 4). With an AAFC of 0, F^. := 0.18; with an AAFC of one, F = 0.27; and with an AAFC of two, F = 0.67. When AAFC was three or more the popu- lation was sustainable at F values up to at least 1.0. Discussion Mortality estimates In most studies where natural mortality has been estimated for shark populations, particularly for use in demographic analysis, indirect estimation meth- ods have been used (e.g. Hoenig and Gruber, 1990; Cortes, 1995; Sminkey and Musick, 1996; Au and Smith, 1997). The results of the current study indi- cate that these indirect methods may provide a wide am res tul trui Jof teal Simpfendorfer: Demographic analysis of Rhizopnonodon taylon 983 Table 3 Sensitivitv of life h istory tables to changes in model structure and age parameters. /,+; m, indicates that age-specific natality was | calculated from the number sharks surviving to the end of age classes; /,m, indicates that age- specific natality was calculated from the numbers surviving at the beginning of classes; 2Mi indicates that natural mortality was doubled for the first age class; IMi indicates that natural mortality was constant for all age classes Definitions of parameters are given in Table 2. Scenario Sensitivity test ^0 G r ; ^2 (years ) F,. I Base case (/^^j m^ , IMj 1.758 2.304 0.271 2.554 0.179 J /,- m, , IMj 3.080 2.304 0.600 1.155 0.600 K /„! m^ . 2Mi 1.003 2.304 0.001 529.2 0.001 L /,m,.,2Mi 1.758 2.304 0.271 2.554 0.271 M Maximum age = 10 1.820 2.488 0.276 2.516 0.183 N Age at maturity = 2 0.035 3.061 0.035 19.82 0.026 range of estimates of natural mortality for a single population. Testing these values of natural mor- tality for R. taylon in life history tables showed that most of the methods produce values of r that would result in a population decreasing even when it was unfished. Such a result is biologically un- reasonable. In contrast, Simpfendorfer (1999) made similar comparisons of indirect estimation methods for the dusky shark (Carcharhinus obscurus) and found that results did not differ greatly between methods. The difficulties in ap- plying many of the indirect methods to R. taylori may relate to its short lifespan and fast growth because all but two of the methods used age or growth parameters to estimate natural mortality. Among the indirect methods for estimating natu- ral mortality, only those of Hoenig (1983) and Gunderson and Dygert (1988) yielded estimates that would allow an unfished population to in- crease. Although the values of natural mortality from the two methods were relatively similar (0.60 and 0.70, respectively), they produced different results when used in the life history tables to cal- culate r, with the Hoenig (1983) estimate produc- ing a much higher value. This difference indicates that the results of demographic analysis are highly sensitive to changes in natural mortality. It is un- likely that the Gunderson and Dygert ( 1988 ) method produced a realistic value of natural mortality be- cause it is based on the assumption that GSl is an index of total reproductive effort. Although this is true for most fish species, it is not the case for pla- cental forms (such as R. taylori) in which much of the energy invested in reproduction is supplied dur- ing development rather than stored in the ovary prior to fertilization. Of the indirect methods used, the Hoenig ( 1983 ) method is therefore likely to be the most realistic estimate of natural mortality for/?, taylori. Fishing mortality (per year) Figure 4 Contour plot of intrinsic rate of population increase (r) as a function of fishing mortality (F) and age at first capture (AAFC) for Rhizoprionodon taylori from northern Australia. Estimates are based on a life table where natural mortality was calculated by a catch curve (scenario H in Table 2). Fish- ing is sustainable at values of r > 0. There are relatively few studies of shark popula- tions in which natural mortality has been estimated directly with either tagging studies (e.g. Grant et al., 1979; Manire and Gruber, 1993) or catch curves (e.g. Cortes and Parsons, 1996). Although direct ap- proaches to determining mortality would be expected to produce more realistic estimates, they must also meet a number of criteria in order to produce accu- rate results. The assumptions for the catch curve technique used in the present paper included con- stant mortality, recruitment, and selectivity over all age classes used in the calculations as well as ran- dom collection of samples from the population. The 984 Fishery Bulletin 97(4), 1999 data on which the calculations for R. taylori were based are likely to have met at least two of these assumptions. Because the area from which samples were collected was subject to little or no fishing, the population size should have remained relatively con- stant and thus fluctuations in recruitment would be minimized. The necessity of excluding the first age class from the catch curve analysis resulted from the selectivity of the gill nets used to collect specimens. For the age classes included in the analysis, the size of individuals was similar because growth rapidly reaches an asymptote, and therefore selectivity was probably reasonably constant. The assumption that mortality was constant was more difficult to test. However, the small size of i?. taylori and the small change in its size after the first year or two of growth (Simpfendorfer, 1993) suggest that all age classes are likely to be subject to similar levels of predation. The estimates of natural mortality for female R. taylori from the catch curve and from the indirect method of Hoenig ( 1983 ) were similar: 0..56 and 0.60, respectively. Such a finding is contrary to that re- ported by Cortes and Parsons ( 1996), who suggested that the Hoenig ( 1983) method would produce lower natural mortality estimates than those from catch curves. The present study suggests that for some species the Hoenig method will produce acceptable results. However, Cortes and Parsons' (1996) caution that demographic studies based solely on a natural mortality estimate from the Hoenig (1983) method should still be heeded because it is clear that the accuracy of this method may vary between species. The most biologically plausible estimates of natu- ral mortality for/?, taylori (0.698 for males and 0.561 for females) are significantly higher than those re- ported so far for many other species of shark. Spe- cies with mortality levels of a similar magnitude are the bonnethead shark (Sphyrna tiburo) {0.625; Cortes and Parsons, 1996) and the Atlantic sharpnose shark {Rhizoprionodon terraenovae) (0.42; Cortes, 1995). Each of these three species (/?. taylori, R. terraenovae, and S. tiburo) are short-lived (6-12 yr). In longer- lived shark species, estimates of natural mortality are much lower. For example, Sminkey and Musick (1996) estimated an M of 0.105 for Carcharhinus pliimbeus, which lives at least 30 years, and Smith and Abramson (1990), Cailliet (1992), and Au and Smith ( 1997 ) estimated that Triakis semifasciata, which lives to 30 years, has an M between 0.139 and 0.15. The small size of i?. taylori probably results in it being subject to high levels of predation throughout life. The small size at birth (220-260 mm) means that the first age class is particularly vulnerable to pre- dation. The results of the sensitivity test, in which natural mortality was doubled for the youngest age class, indicate that this technique, which has previ- ously been used to account for increased predation of young sharks, was not realistic fori?, taylori. This is a result of the high value of natural mortality as- sumed for older age classes. It was not possible to estimate what, if any, increase in predation occurs for the youngest age class of/? . taylori. However, any increase in predation would result in reduced estimates of r. Demographic analysis The most likely demographic results for R. taylori, those based on natural mortality estimated from the catch curve, indicate that the maximum value of r is about 0.27. This value is one of the highest for a spe- cies of shark. Cortes and Parsons (1996) estimated a similar rate (0.272-0.283) for Sphyrna tiburo from Florida with Hoenig's method to estimate natural mortality. Rhizoprionodon taylori and S. tiburo have similar life histories, with short lifespans, rapid growth, early maturity, and high natural mortality. All other published studies of shark demography have estimated intrinsic rate of increase at less than 0.1 (e.g. Hoenig and Gruber. 1990; Cailliet, 1992; Cailliet et al., 1992; Cortes, 1995; Sminkey and Musick, 1996; Au and Smith, 1997; Cortes, 1998). In most cases these species have been longer lived and slower to mature and have exhibited lower natural mortality than /?. taylori or S. tiburo. The exception to this is /?. terraenovae, which grows larger than /?. taylori and is relatively short-lived ( 10 yr), but has a lower reproductive rate (Cortes, 1995). Demographic analysis of/?, taylori is sensitive to the way age-specific reproductive rate is calculated because of this species's short lifespan. Whether age- specific natality is calculated on the basis of the pro- portion of a population surviving at the beginning or the end of an age class leads to a very large differ- ence in r. This effect is magnified by the high natu- ral mortality of/?, taylori. It is therefore important to take account of gestation period in the calculation of reproductive rate. This is less important in longer lived species in which natural mortality is lower and in which there is a relatively small difference between the proportions of a population surviving until the beginning and the end of an age class. This study indicates that accurate age data are required for demographic analysis. The results for /?. taylori showed a high level of sensitivity to changes in age at maturity, but only limited sensitivity to changes in maximum age. Given that the age data on which this research were based have not been validated, there remains some uncertainty about the results. Further work to validate these data (espe- Simpfendorfer: Demographic analysis of Rhizoprionodon taylori 985 cially age at maturity) would decrease uncertainty. However, the length-frequency data supplied by Simpfendorfer ( 1993) support the hypothesis that/?. taylori mature at age one and produce their first lit- ter at age two. Increasing age at first capture for R. taylori in- creased the level of fishing mortality that the popu- lation could sustain. Allowing individuals to remain unfished until theyhad produced one litter (AAFC—2) made fishing mortality levels up to 0.67 sustainable, whereas any level of fishing mortality was sustain- able if individuals were allowed to produce two lit- ters (AAFC=3). In a practical sense, AAFC restric- tions are implemented as size restrictions. On the basis of the growth curves given by Simpfendorfer (1993), size of female R. taylori at the end of each of the first three years is 55. 67, and 71 cm total length (TL), respectively. The fact that size changes only slightly after the first two years makes the use of age restrictions unworkable for older individuals. The gillnet fisherj' in which R. taylori is most of- ten captured in northern Australian waters uses mesh sizes no smaller than 10 cm (stretched mesh). Gill nets of this size rarely catch animals less than 60 cm TL. The age at first capture in this fishery is therefore approximately one year. Although R. taylori less than one year of age are caught by trawlers, this fact was noted by the author only once in observa- tions of research trawls between 1986 and 1992 in the Townsville region. On the basis that age at first capture of i?. taylori in northern Australia is prob- ably one year, demographic analysis indicates that the level of fishing mortality (F^) that the population can sustain is approximately 0.27. However, Ricker (19751 showed that the level of fishing mortality needed to achieve maximum sustainable yield (MSY) is r/2. Thus for R. taylori F^gY = 0.135. Sustainability of fisheries The comparison of demographic assessments of short- lived, fast growing, early maturing elasmobranch species such as R. taylori and S. tiburo, to those of longer-lived, slow growing, late maturing species indicates that they are likely to be able to sustain higher levels of fishing pressure than the latter spe- cies. For example, Sminkey and Musick (1996) re- ported that Carcharhinus plumbeus was most likely to be sustainable when F^ < 0.1, or F^ = 0.25 if a mini- mum size of 178 cm was used. Similarly, Cailliet (1992) reported that for Triakis semifasciata. F^. - 0.084 did not reduce the population, but that fishing mortality at double this level was sustainable only if animals 12 years and older were caught. This con- trasts markedly with the results for R. taylori that the population may be sustainable at f^=0.18, or F^ = 0.27 if age at first capture was one year. It is there- fore possible that at least some species of elasmo- branchs, in particular the shorter-lived, fast grow- ing, early maturing species, are able to sustain lim- ited commercial levels of fishing pressure. The ap- prehensive conclusion of Holden (1974) regarding sustainable fishing of elasmobranch stocks may need to be revised for selected species. However, the real challenge is to manage the development and regula- tion of such elasmobranch fisheries effectively. Acknowledgments This research was based on work carried out at James Cook University, with funding from the University and the Great Barrier Reef Marine Park Authority. Norm Hall provided valuable guidance on the aspects of the analysis and commented upon the manuscript. Tony Hart and two anonymous reviewers also pro- vided valuable comments on the manuscript. Literature cited Au, D. W., and S. E. Smith. 1997. A demogi-aphic method with population density com- pensation for estimating productivity and yield per recruit of the leopard shark {Tiiakis semifasciata). Can. J. Fish. Aquat. Sci. 54:41.5-420. Cailliet, G. M. 1992. Demography of the central California population of the leopard shark [Tnakis semifasciata). Aust. J. Mar. Freshwater Res. 43:183-193. Cailliet, G. M.. H. F. Mollet, G. G. Pittinger, D. Bedford, and L. J. Natanson. 1992. Growth and demography of the Pacific angel shark iSqtiatina californica). based upon tag returns off Cahfornia. Aust. J. Mar. Freshwater Res. 43:1,313-1,330. Casey, J. G., F. J. Mather III, J. M. Mason Jr., and J. Hoenig. 1978. Offshore fisheries of the Middle Atlantic Bight. In H. Clepper led.i. Marine recreational fisheries, vol. 3: pro- ceedings of the second annual marine recreational fisher- ies symposium, p. 107-129. Sport Fishing Institute, Washington. D.C. Cortes, E. 1995. Demographic analysis of the Atlantic sharpnose shark. Rhizoprionodon terraenovae, in the Gulf of Mexico. Fish. Bull. 93:57-66. ' 1998. Demographic analysis as an aid in shark stock as- sessment and management. Fish. Res. 39:199-208. Cortes, E., and G. R. Parsons. 1996. Comparative demography of two populations of the bonnethead shark iSphyrna tiburo). Can. J. Fish. Aquat. Sci. 53:709^717. Grant, C. J., R. O. Sandland, and A. M. Olsen. 1979. Estimation of growth, mortality and yield per recruit of the Australian school shark, Galeorhinus australis 986 Fishery Bulletin 97(4), 1999 (Macleay), from tag recoveries. Aust. J. Mar. Freshwater Res. 30:625 -637. Gunderson, D. R. 1980. Using r-K selection theory to predict natural mortaHty. Can. J. Fish. Aquat. Sci. 37:2,266-2,271. Gunderson, D. R., and P. H. Dygert. 1988. Reproductive effort as a predictor of natural mortal- ity rate. J. Cons. Int. E.xplor Mer 44:200-209. Hoenig, J. M. 1983. Empirical use of longevity data to estimate mortal- ity rates. Fish. Bull. 82( 1 1:898-903. Hoenig, J. M., and S. H. Gruber. 1990. Life-history patterns in the elasmobranchs: implica- tions for fisheries management. In H. L. Pratt Jr.. S. H. Gruber, and T. Taniuchi (eds.i, Elasmobranchs as living resources: advances in the biology, ecology, systematica, and status of the fisheries, p. 1-16. U.S. Dep. Commer , NOAA Tech. Rep. NMFS 90. Holden, M. J. 1968. The rational exploitation of the Scottish-Norwegian stocks of spurdogs iSqualus acanthias L.). Fish. Invest. Ser II2.5(8):l-28. 1974. Problems with the rational exploitation of elasmo- branch populations and some suggested solutions. //; F. R. Harden Jones (ed.). Sea fisheries research, p.ll7- 137. Elek Science, London. 1977. Elasmobranchs. In J. A. Gulland (ed.). Fish popu- lation dynamics, p. 187-215. John Wiley, London. Jensen, A. L. 1996. Beverton and Holt life history invariants result from optimal trade-off of reproduction and survival. Can. J. Fish. Aquat. Sci. 53:820-822. Krebs, C. J. 1985. Ecology: the experimental analysis of distribution and abundance, 3rd ed. Harper and Row, New York, NY, 800 p. Last, P. R., and J. D. Stevens. 1994. Sharks and rays of Australia. CSIRO, Melbourne, 513 p. Manire, C. A., and S. H. Gruber. 1993. A preliminary estimate of natural mortality of age 0 lemon sharks, Negciprion brevirostris. In S. Branstetter (ed.). Conservation biology of elasmobranchs, p. 65- 71. U.S. Dep. Commer, NOAA Tech. Rep, NMFS 115. Musick, J. A., S. Branstetter, and J. A. Colvocoresses. 1993. Trends in shark abundance from 1974 to 1991 for the Chesapeake Bight region of the U.S. mid-Atlantic coast. In S. Branstetter (ed. ), Conservation biology of elas- mobranchs, p. 1-18. U.S. Dep. Commer, NOAA Tech. Rep. NMFS 115. Parker, H. W., and F. C. Stott. 1965. Age, size and vertebral calcification in the basking shark, Cetorhinus maximum (Gunnerus). Zool. Meded. (Leidenl40(34):.305-319. Pauly, D. 1980. On the interrelationship between natural mortality, growth parameters, and mean environmental temperature in 175 fish stocks. J. Cons. Int. Explor Mer 39(2):175- 192. Ricker, W. E. 1975. Computation and interpretation of biological statis- tics offish populations. Bull. Fish Res. Board Can. 191, 382 p. Ripley, W. E. 1946. The biology of the soupfin Galeorlunus zygopterus and biochemical studies of the liver. Fish. Bull. Calif Dep. Fish Game (641:93 p. Simpfendorfer, C. A. 1992. Reproductive strategy of the Australian sharpnose shark, Rhizopnonodon taylori (Elasmobranchii: Carcha- rhinidae), from Cleveland Bay, northern Queensland. Aust. J. Mar Freshwater Res. 43:67-75. 1993. Age and growth of the Australian sharpnose shark, Rhizoprionodon taylori, from north Queensland, Australia. Environ. Biol. Fishes 36:233-241. 1999. Demographic analysis of the dusky shark fishery in south-western Australia. In J. A. Musick (ed.). Life in the slow lane: ecology and conservation of long-lived marine animals, p. 149-160. Am. Fish. Soc. Symp. 23, Bethesda, MD. Simpfendorfer, C. A., and N. E. Milward. 1993. Utilisation of a tropical bay as a nursery area by sharks of the families Carcharhinidae and Sphyrnidae. Environ. Biol. Fishes 37:337-345. Sminkey, T. R., and J. A. Musick. 1996. Demographic analysis of the sandbar shark, Carcharhinus plumbeus, in the western North Atlantic. Fish. Bull. 94:341-347. Smith, S. E., and N. J. Abramson. 1990. Leopard shark Triakis semifasciata distribution, mortality rate, yield and stock replenishment estimates based on a tagging study in San Francisco Bay. Fish. Bull. 88:371-381. Vetter, E. F. 1987. Estimation of natural mortality in fish stocks: a review. Fish. Bull. 86:25-43. i 987 Abstract. -The spotted gully shark, TYiakis megalopterus, was sampled op- portunistically over a 12-year period from catches of shore and ski-boat fish- ermen using hooks and lines. Most specimens (89.6%) were taken from rocky reefs less than 10 m deep, 8% were caught at 11-20 m, and only 2.4% were recorded from waters more than 20 m deep. The reproductive biology of 35 males and 87 females was examined. The spotted gully shark exhibits apla- cental viviparity. Size at 50% maturity for males is ca. 1320 mm total length (TL) and for females ca. 1450 mm TL. Maximum sizes recorded here were 1520 mm TL. for males and 2075 mm TL for females. Gestation appears to last 19-21 months. The female repro- ductive cycle may be 2-3 years, depend- ing on the time between pregnancies. The sex ratio of embryos was found to be 1:1 but the postpartum male:female ratio was 1:2.5. Size at birth was esti- mated to be 420-450 mm TL. The smallest free-swimming individual re- corded was 576 mm TL. Number of embryos per pregnancy ranged between 5 and 15, with a mean of 9.7. A total of 110 stomachs were examined in the feed- ing study. Diet changed with increase in shark size. Small sharks (<1 m) preyed mainly on Cape rock crabs, Plagusia chabrus (78% of mass), whereas sharks of 1-1.4 m preyed largely on Cape rock crabs (48%) and cephalopods (33%). Te- leosts were more important for sharks larger than 1.4 m (54% ); most of these prey were associated with rocky reefs. Reproduction and feeding of spotted gully shark, Triakis megalopterus, off the Eastern Cape, South Africa Malcolm J. Smale Port Elizabeth Museum P.O. Box 13147 Humewood, Port Elizabeth 6013 South Africa E-mail address pemmis g zoo upe ac za Andre J. J. Goosen P.O. Box 1109 Milnerton, Cape Town 7435 South Africa Manuscript accepted 21 January 1999. Fish. Bull. 97: 987-998 (1999). The genus Triakis Miiller and Henle, 1838, commonly called leop- ard shark, comprises small to mod- erately large, harmless, stocky sharks that feed on invertebrates and fish. Triakis is currently re- stricted to five species: Triakis acutipinna Kato, 1968 (sharpfin houndshark); T. rnaculata Kner and Steindachner, 1867 (spotted hound- shark); T. scyllium Miiller and Henle, 1839 (banded houndshark); T. semi- fasciata Girard, 1854 (leopard shark); and T megalopterus (Smith, 1849) (sharptooth houndshark or spotted gully shark; Compagno, 1988). Triakis megalopterus is a stout houndshark with a broadly rounded snout and a large mouth with small, pointed teeth. The head and body are gray or bronze above, usually with numerous small black spots, and white underneath (Bass et al., 1975). It is endemic to southern Af- rica (Compagno et al., 1989), where it occurs from northern Namibia, at about 21°45'S, 13°57'E, to Coffee Bay in Eastern Cape, South Africa, at 31°59'S, 29°09'E. The species is a common inshore bottom-dwelling shark of temperate continental wa- ters, where it is caught by anglers in shallow subtidal waters. It pre- fers rocks and crevices in the shal- lows and is confined to water shal- lower than 50 m along the Cape coast (Bass et al., 1975; Compagno etal., 1989, 1991). Very little has been published on the biology of this species (Bass et al., 1975). Although all five species o{ Triakis are viviparous, they lack yolksac placentas and the embryos obtain nourishment from their yolk sacs (Bass et al., 1975; Castro, 1983; Compagno, 1984; Kusher et al., 1992). Bass et al. (1975) and Com- pagno ( 1984 ) provided some informa- tion on the food habits of T mega- lopterus, noting the presence of crabs {Plagusia chabrus), teleost fishes, and one small shark (Scyliorhinus capen- sis) in stomach contents. At present, there is no scientific information that could guide man- agement decisions on T megalop- terus, with the possibility of a small population size and limited range distribution (Compagno et al. 1989; Goosen, 1997), this species could be vulnerable to overexploitation in in- shore multispecies shark fisheries. Off California, T semifasciata has declined in abundance and conse- quently management measures have been proposed there (Smith and Abramson, 1990; Cailliet, 1992). 988 Fishery Bulletin 97(4), 1999 The aim of this study was to investi- gate the reproductive and feeding biol- ogy of T. megalopterus. Investigations of age and growth will be documented else- where. Improved knowledge of this spe- cies could underpin management strat- egies for this component of the multi- species shark fisheries off South Africa, which' are known to be expanding (Smale, 1997). Materials and methods Specimens were collected over a 12-year period, between February 1984 and Oc- tober 1996, from catches made with hooks and lines by rock, surf, and ski- boat fishermen. All specimens used in this study were collected along the East- ern Cape coastline between Cape St. Francis (34°12'S; 24°52'E) and Coffee Bay (31°59'S; 29°09'E), South Afi^ca (Fig. 1 ). Specimens collected for biological sampling were examined as soon as possible after capture, or were fi-ozen for later study. Sharks were measured to the nearest millimeter and, where possible, weighed with a spring scale accurate to 100 g. Measurements of total length (TL) can vary consid- erably depending on the placement of the caudal fin during measurement ( Branstetter et al., 1987); in this study TL was measured on a horizontal line between perpendiculars, fi-om the tip of the nose to the tip of the tail, with the tail at its maximum extension (Compagno, 1984). Total length is used throughout this paper, un- less otherwise noted. Reproductive information was collected for 35 males and 87 females. Maturity stages largely fol- lowed Bass et al. (1975), with minor modifications (Goosen, 1997). Each specimen was assigned to one of the reproductive stages: embryo, immature, ado- lescent, mature, and, in females, pregnant. The clasper was measured in length from the point of outside insertion in the pelvic fin to the tip of the clasper (CLO); from the point of insertion at the cloaca to the tip of the clasper (CLI); and in width at its thickest point (CBW) (Compagno, 1984). For fe- males, the following data were collected: diameter of the two largest ovarian eggs, width of the oviducal gland, greatest width of the uterus, absence or pres- ence of embryos and uterine eggs, and numbers of embryos and eggs if present. Embryos were measured, sexed, and weighed. The mean length of embryos in each litter was calculated after abnormally developed individuals were ex- cluded. Seasonality of mating, gestation period, and Durban Eastern Cape ♦Xoffee Bay East London '^Pon Elizabeth Cape St Francis 25' 30" 35" Figure 1 Southern Africa showing places mentioned in the text. pupping season were examined by comparing embryo sizes in different months. Hepatosomatic indices (HSI) were calculated from the formula HSI = iLWIBW) X 100, where LW = liver weight in grams; and SW = total body weight in grams. Stomach contents were examined as soon as pos- sible after capture, or frozen for later analysis. Prey were identified to the lowest possible taxon. Excess liquid was drained off and the mass of the remains determined to the nearest 0.1 g on a top-loading pan balance. Bait used to capture the sharks was ex- cluded from analyses. Size measurements used for prey were carapace width (CW) in crabs, mantle length (ML) in cephalo- pods, and total length (TL) in teleosts and all other prey. After whole fish prey were measured, otoliths were removed to verify identification and paired, counted, and measured. Otoliths were used to iden- tify well-digested prey and to estimate their length. Similarly, cephalopod beaks were collected, counted, and measured. Neither formalin nor alcohol was used to store stomach contents, because otoliths exposed to such preservatives become etched or brittle (Smale et al., 1995). Lengths of well-digested cephalopods and teleosts were determined from regressions re- lating beak and otolith lengths to body length ( Smale, 1983; Smale et al., 1993, 1995). Digested otoliths had a chalky eroded appearance and were not measured for use in prey size estimates. Smale and Goosen: Reproduction and feeding of Triakis megalopterus 989 Diet was quantified by 1 ) frequency of occurrence ( ' ; F ). the ratio of stomachs containing a particular prey to stomachs containing any prey, expressed as a per- centage; 2) numerical importance (%N), the number of each prey expressed as a percentage of the total num- ber of prey items; and 3) gi'avimetric importance (%M ), the wet mass of a prey category as a percentage of the total weight of the stomach contents (Hyslop, 1980). By using all three methods of analysis, we avoided bias associated with the use of any one method (Hynes, 1950; Windell, 1968; Hyslop, 1980). No combination of meth- ods was used because this may have resulted in com- liining sources of error ( Berg, 1979 ). Reconstituted prey weights were not used because bias may have been introduced as a result of the different digestion and accumulation rates of fish otoliths and cephalopod beaks ( Smale, 1983). Consequently, the actual wet mass of each item in stomachs was used to investigate prey composition in this study (Smale, 1991). Preliminary analyses revealed considerable differ- ences in prey between sharks of different sizes, and samples were subsequently divided into three arbi- trary size classes: less than <999 mm, 1000-1399 mm, and >1400 mm TL. To investigate the relation between sizes of predator and prey, reconstructed prey lengths and masses were calculated with otolith and beak measurements. Recorded wet masses and carapace widths and lengths had to be used for crus- taceans, and only intact crustaceans were included in the predator-prey size analysis. Common names used for crustaceans are after Holthuis (1991). Results Depth range Material collected in this study was derived from fish- ermen exploiting a wide range of depths. Both shore- based and boat-based catches were sampled, al- though the proportion of effort in different depth ranges was not quantifiable. Nevertheless, T. mega- lopterus was taken mainly from shallow water. Of 125 specimens where collection depth was known, 89.69^ were taken in shallow waters of 10 m and less. Only S% were collected at 11-20 m and 2.4'7f at >20 m. The gi-eatest capture depth recorded in this study was approximately 30 m, at an offshore bank. No obvi- ous size-based habitat segregation was evident in this relatively small sample. Size range and maturation The overall male-to-female ratio in sharks sampled was 1:2.5. The smallest male and female sharks cap- tured were 576 mm and 725 mm, respectively. Fe- males attain larger sizes than males; the largest male sampled was 1520 mm, compared with the largest female of 2075 mm. The clasper length (CLI) of the smallest male was 43 mm and that of the largest (a male of 1402 mm TL) was 239 mm (Fig. 2A). Based on clasper size, degree of calcification, and presence of sperm in the seminal vesicle, maturation in these sharks begins at about 1210 mm and is complete at about 1369 mm (Fig. 2A). Claspers of one specimen of 1345 mm were not quite rigid but sperm presence was noted. The smallest mature male was 1250 mm and the largest immature male was 1196 mm. The claspers of adults measured, on average, 225 mm (range 205-239 mm, « = 16). Increase in the diameter of ovarian eggs indicates the beginning of maturation. Small eggs (<4 mm di- ameter) could be seen in the ovaries of females of 977 mm and larger. Egg diameter increased notice- ably in specimens larger than 1460 mm (Fig. 2B). Mature females had yellow yolk-filled ova larger than 4 mm in diameter. The developing uteri appear as thin strips of translucent tissue with diameters of up to 2.5 mm in females smaller than 1365 mm. As maturation proceeds, the uterus widens first at its posterior end, becoming bottle-shaped at lengths of 1391-1405 mm, with a diameter of 4-16 mm at its widest part (Fig. 2C). One female of 1490 mm TL had a uterus width of 16 mm. All females over 1460 mm, except the previously mentioned individual, had uterus widths wider than 20 mm, and all adult fe- males had uterus widths measuring 20-140 mm ( .V =82.81, n=21). The oviducal gland was difficult to distinguish in the smallest females but usually mea- sured between 3 and 7 mm ( .v=5.64, 7; = 14) in females of less than 1365 mm, and gi-ows little until females reach 1460 mm (Fig. 2D). All females over 1460 mm had oviducal glands wider than 20 mm and all adult females, both pregnant and non-pregnant (resting), had oviducal glands measuring 20-51 mm ( .v=31.7,/!=29). The smallest pregnant female measured 1465 mm TL. Reproduction Of 48 sexually mature females sampled, 81'7f were pregnant. The number of embryos per litter ranged from 5 to 15 ( .v=9.7, n=38). One female had 16 egg cases, but of these only 13 had normal embryos, one was empty, and two contained retarded embryos (length 6.57^ of the next smallest embryo length). Considering the size of the female ( 1650 mm TL). it is possible that litters of up to 16 pups may be re- corded. There is no significant difference in the num- ber of embryos between left and the right uteri 990 Fishery Bulletin 97(4), 1999 o o o o - r-* - c- +J ^ Ul X tj} ■ — « be *■ + ^ ^ ^ ♦^ — CS + + + o o - 'r, * *:*;. ■ 8 c TO § -a *+ + + + ++ • 0°° c o o ■5 a. 0 £ o 7: c o + <- "5 o o c 00 o 1 o 5 S 11 - rn OJ o _' 0- o — re ca cB + • E 2 • SI o .c 3 o ^ c B "2 O ra ^ « 6> o c _4> (/I ■^ o 0 1* o o < .2 Q "5 ^ - § > o o 0 0 0 o - o o « §5 aa-S ft E « - ^ c 0 o 1- , ■ , , , , 1 , 1 , 1 o - o o If c Z. 1 — — T 1 r 1 1 1 u/^ 1 I [ 1 [ ' 1 ' 1 ' 1 ' + • • o o - r- Figure 2 expressed as a p width plotted ag, ■^ "6 o o l^fc o o c c Z •~ • + + + c 0^ •*• i ^J « o 3 o + "3 S; * • ~ ft c • r*^ 0) • 9 o o E E • c u o tr. _1J c c i o o E _4i 3 IS • T3 ^ if o °o o "to O -a < • i "to o o -a < • a; -" i- 2 ^ o o o o 3- c o c t ( < < o - o r — c > 0 a S Jl . O eg cfi c « ra •5 ? "S o c 'Si o - 5 < - o 9^ -c CS 0; o c ■< a, c < o u 2 Relation bet' nan eggs in e il length of fe ' — 1 r r 1 1 1 oc vC TT r J 3' cc -^ - o 1 ' 1 1 1 1 1 1 > 1 ■ oooc ooooc ■O T DO 00 sC -^ D U~J — — — — — « " 00! xil Oouey (UlUl) IjipiM SnJ31| 1 5 § 2 Smale and Goosen: Reproduction and feeding of Thakis megalopterus 991 0 50 100 150 200 250 300 350 400 450 Embryo total length (mm) Figure 3 The relation between embryo mass (EM) and total embryo length ( TL ) was BM = 2.397 x 10-« ■ rL"«i|r=0.99, ;!=275). 1.400 1,500 1,600 1,700 Total length (mm) 1,800 Figure 4 The relation between the number of normal embryos (?2 ) and total length (TL I of pregnant females was n = -21.74 + 0.020 x TL (r=0.52. n=38). (paired t test; P>0.5). Of the 244 embryos that were ■sexed, 122 were male and 122 were female. The relation between wet body mass and total length of embryos was described by a power curve (Fig. 3): W = 2.397 X 10-6 X TL^ 089 [r2=0.994; /7=275] where W TL embryo weight; and embryo total length. Weight of near- term embryos ranged from 300 to 362 g ( .v=331.4g;«=23). During all stages of development, the embryos within a uterus often differed in size; the largest embryos were as much as SO^c longer than the small- est. Greater differences were found in early stages of embryonic development. Larger embryos usually occupy a more posterior position, suggesting that they are older than litter mates in more anterior positions. The embryos in both uteri were at equivalent devel- opmental stages in all the samples. Embryo orienta- tion within the uteri was normally head forward. Only two embryos from one batch lay head backward. Description of the developmental morphology of em- bryos was given by Goosen ( 1997 1. Average observed embryo length in females be- tween 19 May and 25 August was 425 mm ( /? =30 ). A female caught on 19 May had an embryo length range of 416-443 mm ( .v=431 mm;7i = 15), whereas another female, caught on 25 August, had embryos in the length range of 412-440 mm ( .v =425 mm; n =8). Yolk sacs were barely visible in the smaller of these em- bryos, and the largest had closed umbilical scars be- tween the pectoral fins. Denticles and teeth of larger ones had erupted from the skin; one or two black spots were noted on some embryos. Their large size and eruption of teeth and denticles suggested that they had developed approximately to term. No em- bryo examined during this study showed any sign of a placental attachment between embryonic and ma- ternal tissue. Embryos were readily removed from the uterus at all stages of development. Embryos are separated from each other in the uterus by envelop- ing uterine membranes, from which they apparently emerge shortly before birth. The largest embryo measured 443 mm and weighed 341 g. Based on avail- able information, size at birth is estimated at 420- 450 mm; birth probably occurs between May and August. The smallest free swimming individual, 576 mm in length, was taken in August. Careful examination of embryos and uterine eggs showed that some of the uterine eggs failed to de- velop. Of 39 pregnant females, 11 (289f) had some nondeveloping uterine eggs. The greatest number of nondeveloping uterine eggs carried by a single fe- male was 4 out of a total of 11, but most had only one or two. Of these pregnant females, 10 (267r ) contained empty egg cases (no yolk). The largest number of empty egg cases carried by a single female was 12 (out of a total of 121. Occasionally, one embryo in a litter was deformed or retarded in growth. Because of the nondevelopment of a small proportion of uter- ine eggs, the number of developing, apparently nor- mal embryos is a better measure of reproductive out- put than the number of uterine eggs. The relation between number of normal embryos and total length of the mother may be described by a linear regres- sion (Fig. 4): n = -21.74 + 0.02 TL, [n = 381 where /; = litter size; and TL = the total length of the mother in mm. 992 Fishery Bulletin 97(4). 1999 Litter size was found to be significantly correlated with total length of the mother (P<0.01, r=0.52, n=38). The relation between shark length and liver mass was very variable but approximated a power rela- tion (Fig. 5). The maximum recorded weight of fe- males was about three times that of males; varia- tion was higher in mature and pregnant individuals than in immature individuals. To investigate the possible function of the liver as an energy source during pregnancy, female HSI was plotted against mean embryo TL (Fig. 6). Although the trend appeared negative, there was no signifi- cant correlation between mean embryo length and HSI (/ -0.242, P>0.05; n=31). Plotting mean embryo size against month of the year resulted in a scatterplot with two clusters of points but with no clear trend (Fig. 7). Extending the time axis to 24 months and shifting the larger embryos 12 months along the time axis yielded a clear trend. This finding would suggest that there is a ges- tation period of 19-21 mo and that the entire cycle spans approximately two years, excluding any rest- ing period. Of a subsample of 11 mature females taken between May and August, four (369^ ) were rest- ing (they showed no sign of recent pregnancy and had only small ova in the ovary) and the rest were pregnant, five (46%) with pups of intermediate size and two (18'^) with large, near-term pups. Feeding A to 1.200 1 A • •.^* 800- y/ • • 400 - 0^^ * <5 — "^^o"^ ' ' ■ 1 600 800 1,000 1.200 1,400 1,600 1,800 _ Total length (mm) iJ o Juveniles o Adolescem • Mature E > B 3,000 - 2.000 - >0< ' X V o • X V X X /^ •^. X / ^ 1.000 - ryx X • /f mx yb x- ^ X 0 ■ 0-^333'^'''^^ 600 1,100 1,600 Total length (mm) o Juvenile • Adolescent • Mature x Pregnant Figure 5 (A) The relation between liver mass (LM) and total length of male T. megalopterus was LM = 9.517 x IQ-io x TL 3 "94 (r=0.83, n=28). (B) The relation between liver mass (LM) and total length of female T. megalopterus was LM = 1.964 x lO-s x TL'' "52 lr=0.86, n=73). tal of 1 10 stomachs were examined, consisting of 34 males (576-1520 mm TL) and 76 females (660-1746 mm TL). Females had a higher percentage of empty stomachs (20.5%) than males (12.1%). Preliminary analysis indi- cated a change in diet with predator size. This was investigated by grouping the data into three arbitrary size classes: smaller than >999 mm TL, 1000-1399 mm TL, and >1400 mm TL and longer. Prey taken clearly changes with increasing size ( Fig. 8 ). Although invertebrates were important initially, larger sharks took more vertebrates (Table 1). Small sharks fed almost entirely on crabs (Fig. 8). The Cape rock crab, Plagusia chabrus, dominated in the smaller two size classes of shark, in terms of mass (78.3% and 48.5%, respectively; Table 1). Larger crustacean taxa appeared in the diet of larger sharks. Cape slipper lobster, Scyl- larides elisabethae, appeared in the second size class and Cape rock lobster, Jasus lalandii, and scalloped spiny lobster. Pan- 16 T 14 - 12 ■ I 8 - 6 - 4 ■ • ^"^"TVT ^ ^ •• • •• • • ( ) 100 200 300 400 Embryo total length (mm) Figure 6 The relation between hepatosomatic index (HSI) of pregnant females and the mean length iTLi of normally formed embryos in the litter was HSI = 9. .376 - 0.0068 TL (r=0.242. n=31). Smale and Goosen: Reproduction and feeding of Triakis mega/opferus 993 ulirus homarus, appeared in the largest size class. Although crustaceans were important prey of the large sharks, they were less dominant in the diet. Teleosts became more important prey for larger sharks. They included representatives from at least 10 families and 14 species (Table 1). They dominated the prey of large sharks by mass (53.9%) and by num- ber i.477c), whereas they contributed only 10.3% and 14.2% by mass for the medium and small-size classes, respectively (Fig. 8). Only three species offish were eaten by the smallest sharks, of which barred fingerfin, Cheilodactylus pixi, was the most impor- tant. Six species were taken by the medium-size class, and twotone fingerfin (Chlrodactylus brachydactyhis) was the most important. The diet of the large-size class was dominated by several species, viz. seacatfish (Galeichthys sp. ), cob (Ar-gyrosomiis inodorus). red tjor- tjor (Pagellus bellottii natalensis), and blue hottentot (Pachymetopon aeneum). The two largest teleost spe- cies ingested, based on recalculated total length, were a 336-mm sand steenbras (Lithognathus mormyrus) and a 334-mm seacatfish (Galeichthys sp.). Both were taken by sharks longer than 1500 mm TL. Cephalopods were the second most important for- age category for medium-size sharks (Fig. 8). The common octopus [Octopus vulgaris) and the squid Loligo vulgaris reynaudii were the most important species, making up 25.5% and 7.8% by mass, respec- tively. Cephalopods were minor prey for the largest class of sharks (4.6% by mass). £ 400 -I j= ^ 300 _a> 1 200 - o E 0 •• • ••• "•"Bassetal-v 1975 I I I I I I I I I I I I I I ■■ r 1 ' I * I I 0 3 6 9 12 15 18 21 24 Months Figure 7 Mean size of normal embryos in each lit- ter plotted against month of the year. The arrow indicates a shift of large lit- ters (empty circles) to the same months one year later Elasmobranchs were relatively unimportant prey of spotted gully sharks. One stomach had the remains of a lesser sandshark (Rhinobatos sp.), another con- tained remains of a brown catshark (Haploblepharus fuscus ), and two contained catshark eggs. The occur- rence of anomalous food items was very rare. Ined- ible remains included a bryophyte (one stomach) and an unidentified mussel (one stomach). In addition to ontogenetc variation in prey, prey size increased with growth (Fig. 9). Cape rock crab {Plagusia chabrus) was an important prey for sharks <999 " 1.000-1.399 Crustaceans Cephalopods Teleosts Prey taxa Elasmobranchs Miscellaneous Figure 8 Percentage mass of different taxa in the diet for different size classes of T! megalopterus. 994 Fishery Bulletin 97(4), 1999 Table 1 Prey composition of three size groups of 7^ megalopterus Toni the Eastern Cape. < 999 mm TL 1,000 -1,399 mmTL > 1,400 mm TL %F %N %M %F %N %M %F %N %M Crustaceans Jitergatis roseus 2.38 1 1.63 Ovalipes trimaculatus 4 1.89 0.27 4 1.39 0.02 9.52 5 1.83 Plagusia chabrus 76 69.81 78.29 60 50 48.48 30,95 16 16.26 Jasus lalandii 2.38 1 6.65 Panulirus homarus 4.76 2 4.78 Scyllarides elisabethae 8 2.78 1.72 11.90 6 4.65 Caridea 8 3.77 0.06 2.38 1 0.11 Unidentified crab spp. 20 13.21 7.18 16 8.33 2.45 9.52 5 0.86 Molluscs Cephalopods Loligo vulgaris reynaudii 8 2.78 7.83 2.38 1 0.96 Octopus vulgaris 24 13.89 25.48 7.14 3 1.72 Sepia sp. 2.38 1 0.01 Gastropods Ha Hot is midae 4 1.39 2.51 2.38 1 1.90 Teleosts Ariidae Galeichthys sp. 11.90 6 13.10 Carangidae Trachurus trachurus capensis 4 1,39 2.32 Cheilodactylidae Chedodactylus pixi 8 5.66 13 4 1.39 0.07 2.38 1 0.45 Chirodactylus brachydactylus 4 1.39 3.98 Gobiesocidae Chorisochismus dentex 2.38 1 1 Haemulidae Pomadasys olivaceum 4 3.77 0.79 Mugilidae Liza richardsonii 2.38 1 0.94 Platycephalidae 2.38 1 0.01 Pomatomidae Pomatomus saltatrix 7.14 2 0.70 Sciaenidae Argyrosomus inodorus 4 1.39 0.93 4.76 4 8.50 Sparidae 4,76 2 0.01 Diplodus sargus capensis 4 1.89 0.40 4 1.39 0.50 2.38 1 0.10 Lithognathus mormyrus 2.38 1 3.37 Pachymetapon aeneum 4.76 6 4.65 Pagellus bellottii natalensis 4 1.39 1.92 4.76 5 5.25 Sarpa salpa 2.38 2 2.76 Unidentified teleost remains 16 5.56 0,61 30.95 14 13.02 Elasmobranchs 2.38 1 0.18 Scyliorhinidae 2.38 1 0.16 Haploblepharus fuscus 2.38 1 0.94 H. fuscus (egg cases) 4 2.78 0.28 2.38 2 0.31 Rhinidae Rhinobatos sp. 2.38 1 2.65 Skate sp. (egg case) 2.38 1 0.31 Miscellaneous Bryophyte 4.76 2 0.02 Unidentified mussel 4 1.39 0.86 Tapeworm 4 1.39 0.03 Unidentified material 2.38 1 0.20 Total' 25 53 1092.64 25 72 3,014.90 42 100 4,898.95 ' Totals are numbers of stomachs with prey, number of prey items and prey w • I Smale and Goosen: Reproduction and feeding of Trlakls megalopterus 995 600 H A 3 500 - i^ 400 - 1 300- £- 200- £ 100 - ■ A 500 1,000 1,500 Total length (mm) o Crustaceans ■ Cephalopods a Teleosts Figure 9 Calculated original prey mass plotted against T. megalopterus length, showing that the size of prey taken increases with predator size. of a wide range of sizes, and there was a significant linear relation between crab carapace width and to- tal length of sharks (r=0.37, P<0.05, n=39; Fig. lOA). Limited numbers of teleosts made analysis difficult. Combined data for two species (Cheilodactylus pixi and Diplodus sargus) showed a significant relation between the lengths of prey and sharks (r=0.97, P<0.01; n=l; Fig. lOB). The sample size of other prey species was too small to make similar analyses. Discussion Depth range This study confirmed earlier reports that T. mega- lopterus prefers shallow rocky reefs (Bass et al., 1975; Compagno et al., 1989, 1991). They were found very rarely in deep waters, despite extensive sampling in deep waters for a variety of other sharks, including the closely related Mustelus species (Smale, 1991; Smale and Compagno, 1997). The nature of the samples precluded precise determination of intraspe- cific habitat choice, but there were no obvious differ- ences in habitat choice by sharks of different sizes. The skewed sex ratio, which needs further investigation, may be due to sampling bias or social factors. Reproduction In T. megalopterus, eggs produced by the single func- tional ovary pass through the oviducal gland, where they are probably fertilized before they are enclosed in the membranous egg case. Embryonic development -is ovoviviparous, embryos receiving nourishment from the yolk sac. However, some egg cases were slightly adherent and others intimately connected 600 o Juvenile 800 1.600 1,000 1,200 1,400 Predator length (mm) Adolescent • Mature * Pregnant E £ 300 ■5 250 J 200 - 1 150 - i 100 S 50 B 600 800 1,000 1.200 1.400 1,600 Predator length (mm) • Cheilodacn'ltis pixi A Diplodus sargus capensis Figure 10 Relation between prey size and predator length. (A) Plagusia chabrus carapace width (CW) plotted against predator length. CW = 25.1 + 0.014 x TL (r=0.37, f!=39). (B) Total lengths of two teleosts plotted against total length of the predator The linear relation had the form TL (prey) = -41.94 + 0.176 ■ TL (predator) (r=0. 97, n=l). to the uterus wall. There was no evidence of the yolk sac forming a placental attachment at later stages of development, supporting Compagno's ( 1984) sum- mary of the genus. The size at which 50% of male sharks are mature is equivalent to mean size at maturity (Lenanton et al., 1990). Data from the present study indicated that males mature at approximately 1320 mm. Compagno (1984) estimated that males mature at 1300-1400 mm. On the basis of width of the oviducal gland and the diameter of the uterus and eggs, females begin to mature at about 1391 mm, and 507f maturity is achieved at about 1450 mm; all females larger than 1500 mm are mature. Compagno ( 1984) estimated that females mature between 1400 and 1500 mm and re- ported mature females of 1400-1740 mm. In this study, 50^^ maturity of female sharks is attained at about 1450 mm, which represents 70% of the maximum size ob- served, 2075 mm. This size falls within the range of 60%-90% noted by Holden and Raitt ( 1974). The small- est pregnant female measured 1465 mm. Gestation in sharks is usually 10-12 mo and pro- longed gestation periods are apparently rare, al- though Squalus acanthias has a gestation period of 996 Fishery Bulletin 97(4), 1999 22-24 mo (Wourms, 1977; Nammack et al., 1985; Hanchet, 1988; Wourms and Demski, 1993; Wourms, 1994). The gestation of frilled sharks, Chlamy- doselachus anguineus, may be as long as three and a half years (Tanaka et al., 1990). The proposed gesta- tion of about 20 mo in T. megalopterus is longer than the estimated 12 mo in Triakis seinifaciata (Castro, 1983; Talent, 1985; Smith and Abramson, 1990). If the 20'-mo estimate is correct, any females that mate shortly after parturition would have a resting period of two to three months. Some mature but nonpreg- nant females, however, show no sign of ovarian egg growth and development at the same time that oth- ers have large ovarian eggs or uterine eggs (or both) and small fetuses. These resting females represented 36% of the sample of 11 mature females taken be- tween May and August. These females probably skip a year and have an extended reproductive break of at least 12-15 mo, which would prolong the cycle to at least 3 years for some individuals. On the other hand, some pregnant females about to undergo par- turition have large ovarian eggs, and these individu- als may mate within a few months and apparently forego the extended resting period. On the basis of embryo growth rates presented above, mating and fertilization probably occur from about October to early December Females carry term embryos of 422-440 mm (largest embryos) between the last week of May and the last week of August, which would approximate the time of parturition. Compagno's ( 1984 ) estimate of size at birth was much smaller (300-320 mm) than that of our findings. From observations of embryonic development, it was noted that during the late stages of embryo development (from ca. 400 mm), both external and internal yolks were absent. The small amount of yellow substance in the spiral valve was thought not to be an internal yolk reserve, but the area of absorption. It was also noted that each embryo was contained in a soft egg case with ca. 500 mL of fluid (Goosen, 1997). Wourms and Demski (1993) noted that in Squalus acanthias, several months into the 22-mo gestation period, em- bryos can ionoregulate and osmoregulate in a uter- ine solution resembling seawater This might be the case for T. megalopterus and could contribute to near- term embryo nourishment until parturition occurs. The liver mass of a shark is a good index of the shark's condition (Springer, 1960). In our study it was found to fluctuate widely, particularly after matura- tion. Seasonal variation in HSI of the lesser sandshark (Rhinobatos annulatus) has been attrib- uted to fluctuations in lipid content of the liver, which has been correlated with reproductive condition in females (Rossouw, 1987). Similar changes in HSI have been recorded in pregnant females of three Mustelus species (King, 1984; Smale and Compagno, 1997). Although low HSI values were found in preg- nant females with near-term embryos in this study, the relation was not statistically significant, possi- bly suggesting that HSI fluctuates with other physi- ological factors. These observations were based on a small sample size and need further investigation. Feeding This study showed that T. megalopterus changes diet with growth, which suggests that it selects prey of suitable size. Because there is no evidence to date of marked habitat change with growth, and all sizes of shark do not have the same prey, it is unlikely that they are feeding opportunistically, even if they favor abundant species. Opportunistic feeding sensu Wetherbee et al. (1990) implies that stomach con- tents are varied but of composition and abundance similar to those of prey in the environment. Smale ( 1996 ) noted that the term "opportunistic" should be used with caution because it is difficult to distinguish between abundance and availability. Ontogenetic variation in feeding of T. megalopterus appears to be attributable to broadening the diet to include more energetically rewarding species (e.g. teleosts), al- though benthic prey are still taken. Spotted gully sharks fed primarily on the crab Plagusia chabrus; as sharks grew, larger individu- als were taken. This finding conforms with those where large crabs were taken by larger leopard sharks (T. semifasciata,Ta\ent 1976). Although spot- ted gully sharks are most abundant near reefs, they may also hunt over sandy areas to exploit the crab Ovalipes trimaculatus, which prefers sandy sub- strates. Spotted gully sharks preyed largely on noc- turnally active lobsters (Paterson, 1969; Smale, 1978; Zoutendyk, 1988) and crabs (Brown, 1961; Warner, 1977), suggesting that they are nocturnal hunters that take crustaceans as they emerge. Such activity may explain why anglers catch them more frequently at night. As the sharks increase in size, they use a much wider variety of prey groups, including teleosts. Therefore, with growth the spotted gully shark is able to attack and ingest larger prey species. Nocturnally hunting sharks may be able to take some teleosts when they are less active or resting at night. Elas- mobranchs became increasingly important prey with growth, a finding that supports previous observations of congeners taking elasmobranchs (Bass etal., 1975; Russo, 1975; Talent, 1976; Compagno, 1984). Octopus vulgaris was the most common cephalo- pod prey It inhabits reefs (Smale and Buchan, 1981; Roper et al., 1984) and emerges from its den to hunt Smale and Goosen: Reproduction and feeding of Triakis mega/opterus 997 between dusk and dawn (Smale and Buchan, 1981), making it vulnerable to nocturnal predators. Squid iLoligo vulgaris reynaudii) is common throughout the Agulhas Bank (the continental shelf between Cape Town and Port Elizabeth) but become concentrated inshore in spring and early summer, when it spawns (Sauer and Smale, 1991; Augustyn et al., 1992). Egg laying is concentrated on the bottom during the day in waters 10-50 m deep (Augustyn, 1990; Sauer et al., 1997). Squid are thought to be especially vulner- able to predation during mating and spawning (Smale, 1991; 1996). Sauer and Smale (1991) re- corded T. megalopterus as one of the squid predators in the vicinity of spawning aggregations, which il- lustrates its ability to exploit a superabundant prey resource by day, even though it normally appears to hunt at night. In conclusion, it is evident that the habitat prefer- ences of gully sharks make them vulnerable to ex- ploitation by inshore fishermen. Their life history traits — large size at maturity, prolonged gestation period, and relatively small litter size — suggest that this species is unsuitable for sustained harvesting by either recreational or commercial fishing. Acknowledgments We are grateful to fishermen who donated specimens for research and thank those colleagues who helped with specimen collection in the field. We thank D. Baird of the University of Port Elizabeth and anony- mous referees for criticisms of an earlier draft. The Foundation for Research Development and the South African Network for Coastal and Oceanic Research (Marine and Coastal Resource Programme) provided financial support. Literature cited Augustyn, C. J. 1990. Biological studies on the chokker squid Loligo vul- garis reynaudii (Cephalopoda; Myopsida) on spawning grounds off the south-east coast of South Africa. S. Afr. J. Mar Sci, 9:11-26. Augustyn, C. J., M. R. Lipinski, and W. H. H. Sauer. 1992. Can the Loligo squid fishery be managed effectively? A synthesis of research on Loligo vulgaris reynaudii. S. Afr. J. Mar Sci. 12:903-918. Bass, A. J., J. D. D. Aubrey, and N. Kistnasamy. 1975. Sharks of the east coast of southern Africa. III. The families Carcharhinidae (excluding Mustelus and Car- charhinus) and Sphyrnidae. Oceanogr. Res. Inst. (Durban) Invest. Rep. 38, 100 p. Berg, J. 1979. Discussion of methods of investigating the food of fishes, with reference to a preliminary study of the prey of Gobiusculus flavescens (Gobiidae). Mar Biol. 50: 263-273. Branstetter, S., J. A. Musick, and J. A. Colvocoresses. 1987. A comparison of the age and growth of the tiger shark, Galeocerdo cuvieri, from off Virginia and from the north- western Gulf of Mexico. Fish. Bull. 85:269-279. Brown, F. A. 1961. Physiological rhythms, /n T. H. Waterman (ed.). The physiology of Crustacea, vol. II, p. 401-426. Academic Press. New York. NY. Cailliet, G. M. 1992. Demography of the central California population of the leopard shark ( Triakis semifasciata ). In J. G. Pepperell (ed.), Sharks: biology and fisheries. Aust. J. Mar. Fresh- water Res. 43:183-193. Castro, J. I. 1983. The sharks of North American waters. Texas A&M Univ Press, College Station, TX, 180 p. Compagno, L. J. V. 1984. FAO species catalogue. 4. Sharks of the world: an annotated and illustrated catalogue of shark species known todate. (2)Carcharhiniformes. FAO Fish Synop. 125:251- 655. 1988. Sharks of the order Carcharhiniformes. Princeton Univ Press, NJ. 486 p. Compagno, L. J. V., D. A. Ebert, and P. D. Cowley. 1991. Distribution of offshore demersal cartilaginous fish (Class Chondrichthyes) off the west coast of southern Af- rica, with notes on their systematics. S. Afr J. Mar. Sci. 11:43-139. Compagno, L. J. V., D. A. Ebert, and M. J. Smale. 1989. Guide to the sharks and rays of southern Africa. Struik, Cape Town, 160 p. Goosen, A. J. J. 1997. The reproduction, age and growth of the spotted gully shark, Triakis megalopterus. off the Eastern Cape coast. M.Sc. thesis, Univ. Port Elizabeth, South Africa, 97 p. Hanchet, S. 1988. Reproductive biology of Squalus acanthias from the east coast. South Island. New Zealand. N.Z. J. Mar Fresh- water Res. 22:537-549. Holden, M. J., and D. F. S. Raitt. 1974. Manual of fisheries science. Part 2. Methods of re- source investigation and their application. FAO Fish. Tech. Paper 115, rev 1, 214 p. Holthuis, L. B. 1991. FAO species catalogue. 13. Marine lobsters of the world: an annotated and illustrated catalogue of species of interest to fisheries known to date. FAO Fish Synop. 125, 292 p. Hynes, H. B. N. 1950. The food of fresh-water sticklebacks iGasterosteus aculeatus and Pygosteus pungitius), with a review of meth- ods used in studies of the food of fishes. J. Anim, Ecol. 19:36-58. Hyslop, E. J. 1980. Stomach content analysis — a review of methods and their application. J. Fish Biol. 17:411-429. King, K. J. 1984. Changes in condition of mature female rig (Mustelus lenticulatus) from Golden Bay in relation to seasonal in- shore migrations. N.Z. .J. Mar. Freshwater Res. 18: 20-27. Kusher. D. I., S. E. Smith, and G. M. Cailliet. 1992. Validated age and growth of the leopard shark, Triakis semifasciata, with comments on reproduction. Env. Biol. Fish. 35:187-203. 998 Fishery Bulletin 97(4). 1999 Lenanton, R. C. J., D. I. Heald, M. Platell, M. Cliff, and J. Shaw. 1990. Aspects of the reproductive biology of the gummy shark, Mustelus antarcticus Giinther, from waters off the south coast of Western Australia. Aust. J. Mar. Freshwa- ter Res. 41:807-822. Nammack, M. F., J. A. Musick, and J. A. Colvocoresses. 1985. Life history of spiny dogfish off the Northeastern United States. Trans. Am. Fish. Soc. 114:367-376. Patersqp, N. F. 1969. The behaviour of captive Cape rock lobsters, Jaxiis lalandii ( H. Milne Edwards). Ann. S. Afr. Mus. 52:225-264. Roper, C. F. E, M. J. Sweeney, and C. E. Nauen. 1984. FAO species catalogue. 3. Cephalopods of the world: an annotated and illustrated catalogue of species of inter- est to fisheries. FAO Fish Synop. 125:221-212. Rossouw, G. J. 1987. Function of the liver and hepatic lipids of the lesser sand shark. Rhinohatns aniiulatiis (Muller & Henle). Comp. Biochem. Physiol. B 86:785-790. Russo, R. A. 1975. Observations on the food habits of leopard sharks (Triakis semifasciata ) and brown smoothhounds iMustelus hcnlei). Cal. Fish Game. 61:95-103. Sauer, W. H. H., and M. J. Smale. 1991. Predation patterns on the inshore spawning grounds of the squid Loligo vulgai'is reynaudii (Cephalopoda: Loliginidae) off the south-eastern Cape, South Africa. S. Afr. J. Mar. .Sci. 11:513-523. Sauer, W. H. H, M. J. Roberts, M. R. Lipinski, M. J. Smale, R. T. Hanlon, D. M. Webber, and R. K. O'Dor. 1997. Choreography of the squid's "nuptial dance." Biol. Bull. 192:203-207. Smale, M. J. 1978. Migration, growth and feeding in the Natal rock lob- ster. Panulirus liomanis I Linnaeus). Oceanogr. Res. Inst. • Durban) Invest. Rep. 47, 56 p. 1983. Resource partitioning by top predatory teleosts in eastern Cape coastal waters (South Africa). Ph.D. diss., Rhodes Univ., South Africa, 275 p. 1991. Occurrence and feeding of three shark species. Carcharhinus brachyurus, C. obscurus and Sphyrna zygaena, on the Eastern Cape coast, South Africa. S. Afr. •J. Mar. Sci. 11:31-42. 1996. Cephalopods as prey IV. Fishes. Phil. Trans. R. Soc. Lond. B Biol. Sci. 351:1,067-1,081. 1997. Trade in sharks and shark products in South Africa. In N.T. Marshall and R. Barnett (eds.). Trade re- view: the trade of sharks and shark products in the west- ern Indian and south-east Atlantic oceans, p. 80- 100. TRAFFIC East/Southern Africa, Nairobi. Smale, M. J., and P. R. Buchan. 1981. Biology of Octopus vulgaris off the East coast of South Africa. Mar. Biol. 65:1-12. Smale, M. J., M. R Clarke, N. T. W. Wages, and M. A. C. Roeleveld. 1993. Octopod beak identification — resolution at a regional level (Cephalopoda, Octopoda: southern Africa I. S. Afr. J. Mar. Sci. 13:269-293. Smale, M. J., and L. J. V. Compagno. 1997. Life history and diet of two southern African smooth- hound sharks, Mustelus mustelus (Linnaeus, 1758) and Mustelus palumbes Smith, 1957 (Pisces: Triakidae). S. Afr. J. Mar. Sci. 18:229-248. Smale, M. J., G. Watson, and T. Hecht. 1995. Otolith atlas of southern African marine fishes. Ichthyol. Monogr. J. L. B. Smith Inst. Ichthyol., 253 p. + 149 plates. Smith, S. E., and N. J. Abramson. 1990. Leopard shark Ti-iakis semifasciata distribution, mortal- ity rate, yield, stock and replenishment estimates based on a tagging study in San Francisco Bay Fish. Bull. 88:371-381. Springer, S. 1960. Natural history of the sandbar shark Eiilamia mil- bcrti. U.S. Fish Wildlf Serv, Fish. Bull. 61:1-38. Talent, L. G. 1976. Food habits of the leopard shark, Triakis semifasciata, in Elkhorn Slough, Monterey Bay, California. Cal. Fish Game 62:286-298. 1985. The occurrence, seasonal distribution, and reproduc- tive condition of elasmobranchs fishes in Elkhorn Slough, California. Cal. Fish Game 71:210-219. Tanaka, S., Y. Shiobara, S. Hioki, H. Abe, G. Nishi, K. Yano, and K. Suzuki. 1990. The reproductive biology of the frilled shark, Clilamy- t/oselaclius anguineus, from Suruga Bay, Japan. -Jpn. J. Ichthyol. 37:102-120. Warner, G. F. 1977. The biology of crabs. EIek .Science, London, 197 p. Wetherbee, B. M., S. H. Gruber, and E. Cortes. 1990. Diet, feeding habits, digestion, and consumption in sharks, with special reference to the Lemon shark, Negaprion hrevirostris. In H. L. Pratt, S. H. Gruber, and T. Taniuchi (eds), Elasmobranchs as living resources: ad- vances in the biologv, ecology, systematics, and the status of the fisheries, p. 29-47. U.S. Dep. Commer, NOAATech. Rep. NMFS 90. Windell, J. T. 1968. Food analysis and rate of digestion, //i W. E. Ricker (ed.). Methods for assessment offish production in fresh waters, p. 197-203. Blackwell, Oxford. Wourms, J. P. 1977. Reproduction and development in Chondrichthyan fishes. Am. Zool. 17:379-410. 1994. The challenges of piscine viviparitv Isr. -J. Zool. 40: 551-568. Wourms, J. P., and L. S. Demski. 1993. The reproduction and development of sharks, skates, rays and ratfishes: introduction, history, overview, and fu- ture prospects. Env. Biol. Fishes 38: 7-21. Zoutendyk, P. 1988. Feeding, defaecation and absorption efficiency in the Cape rock lobster Jasus lalandii. S. Afr J. Mar Sci. 6:59-65. \ 999 Abstract.— Continuous observations in a large research aquarium (121 kL) over a 14-week period were combined with field collection in the Navesink River estuary, New Jersey, to explore the behavior of winter flounder [Pseudo- pteuronectes americanus) during and after the spawning season. Ten males and ten females held in the aquarium spawned over a 60-day period, with an average of 40 spawns per female and 147 spawns per male. Males initiated all observed spawning events, which occurred throughout the night, but pri- marily between sunset and midnight. Spawning by one pair frequently elic- ited sudden convergence and spawning by secondary males (up to six individu- als!; consequently, strictly paired spawning was uncommon (22'?^ of events). Males and females were almost entirely nocturnal during the reproduc- tive season but became increasingly di- urnal during the postspawning season. Males and females arrived in the es- tuary in ripe spawning condition; fish with high gonadosomatic indices were collected during most of February and March 1997. All of the ripe females were >20 cm in total length and at least two years old, whereas ripe males mea- suring 10-15 cm were common. Ripe males were found throughout the sys- tem whereas ripe females were concen- trated in the middle reach of the estu- ary Field collections revealed that fe- males began feeding, primarily on ampeliscid amphipods and on siphons of the clam Mya arenaria earlier in the season than males. Males in the labo- ratory also began feeding late in the season, after most spawning had ended. Field and laboratory results combine to indicate that male spawning strategy is adapted to maximize numbers of eggs fertilized. There is probably high ge- netic diversity in the offspring from any one female owing to frequent spawn- ing and to multiple males participat- ing in individual spawning events. Behavior of winter flounder, Pseudopleuronectes americanus, during the reproductive season: laboratory and field observations on spawning, feeding, and locomotion Allan W. Stoner Allen J. Bejda John P. Manderson Beth A. Phelan Linda L. Stehlik Jeffrey P. Pessutti Northeast Fisheries Science Center National Marine Fisheries Service, NOAA 74 Magruder Road, Highlands. New Jersey 07732 E-mail address (for A W Stoner) al stoner a noaa gov Manuscript accepted 24 March 1999 Fish. Bull. 97: 999-1016 (1999). Winter flounder, Pseudopleuronectes americanus. is an important commer- cial and recreational fishery species in estuarine and continental shelf habitats along the Atlantic coast of North America from Labrador to Georgia (Scott and Scott, 1988). A large number of investigations have been devoted to this species and its life history is relatively well known (Klein-MacPhee, 1978; Able and Fahay, 1998 ). Winter flounder popu- lations include several substocks that mix during summer residence on the shelf, but adults return to natal estuaries to spawn ( Saila, 1961; Poole, 1966; Howe and Coates, 1975; Pierce and Howe, 1977; Phelan. 1992). In the Mid-Atlantic Bight, adults make seasonal migrations into the estuaries of New York and New Jersey (Perlmutter, 1947; Phelan, 1992) where they spawn in shallow inshore waters during late winter (Scarlett and Allen, 1992). Winter flounder in the mid-Atlan- tic region are believed to mature at 2-3 years, at 20-25 cm total length (Perlmutter, 1947, Danila, 1978; Witherell and Burnett, 1993). Con- siderable attention has been given to seasonal movement (Perlmutter, 1947; Saila, 1961; Howe and Coates, 1975; Phelan, 1992) and to fecun- dity and reproductive state in wild populations (Kennedy and Steele, 1971; Danila, 1978; Burton and Idler, 1984; Nelson et al., 1991); however, information on spawning behavior is limited to observations made by Breder (1922) of fish in small laboratory tanks. In this investigation we expanded Breder's ( 1922 ) description of spawn- ing and courtship in winter flounder by making continuous observations on a population of winter flounder in a large research aquarium for a 14-week period during and after the reproductive period. We determined the total number of spawning events for males and females in the population and examined seasonal and diurnal variation in spawning, feeding, and locomotory activity. On the basis of field collection we de- scribed population structure, go- nadal development, feeding, and localized movements by a wild population of winter flounder in an 1000 Fishery Bulletin 97(4), 1999 estuarine spawning ground during winter and spring 1997. Methods Laboratory observations Experrmental animals Winter flounder were col- lected in nearshore waters between Rockaway Point and Coney Island, New York, on 27 January 1997 with a 9.2-m otter trawl fished from RV Gloria Michelle. Flounder were held in live cars with flow- through seawater and transported in large ice chests to the laboratory. Mature ripe males were identified by the extrusion of sperm under light pressure ap- plied to the body Females were identified by the pres- ence of full ovaries distending the body, occupying a large portion of the body cavity, and extending al- most to the caudal peduncle. Ten males and ten fe- males, each greater than 200 mm in TL (total length), were marked with Peterson tags (13 mm diameter) before release into the research aquarium. To pro- vide visual identification of gender on videotape, males and females were marked with orange and white tags, respectively. These colors were distin- guishable with low-light video cameras during both day and night. At the beginning of the experiment the mean total length of male flounders was 29.8 cm (SD=3.2, range=24.3-34.4 cm). Mean length of fe- males was 28.5 cm (SD=2.6, range=23. 5-30.8 cm). Four fish (two males and two females) died near the end of the 14-week observational period; these were not replaced. The fish were remeasured for total length at the end of the experiment to estimate growth. Research aquarium Observations of the winter flounder were made in the research aquarium at the James J. Howard Marine Sciences Laboratory at Sandy Hook, New Jersey. The aquarium (121 kL vol- ume) is an oval 10.6 m long, 4.5 m wide, and 3 m deep, with eight rectangular windows (0.7 m wide and 1.2 m high), one in each end and three along each side. There is a recirculating system so that 10*)^ of the water is replaced each week. Seawater upwells through coarse sand which covers the bottom of the aquarium (46 m^) to a depth of 40 cm, then exits through drains at the top of the tank. For this experiment, photoperiod in the aquarium room, which was controlled 'by a computer-driven bank of fluorescent lamps (Duro-Test, Vita-Lite: model CRI91; 5500°K), was programmed to follow the natural sequence at the location of the labora- tory (40'='28'N, 74°00'Wi, with a 13.5-h dark period at the end of January, when fish were first introduced into the system, and a 10-h dark period at the end of the study in mid-May. During nighttime hours, light intensity on the sediment surface was 0.17 lux (1.98 X 10"'' niEinsteins PAR [photosynthetic active radia- tion]). Light intensity increased at the time of twi- light to 2.0 lux (2.92 X 10-5 mEinsteins PAR) for 20 min, followed by a natural rise in intensity to 206 lux (0.0027 mEinsteins PAR) at midday. Return to night- time conditions followed an exact reverse pattern. Water temperature was held at 4.0°C (±0.5°C) from the beginning of the observations until 27 March, when a cooling system malfunction allowed tempera- ture to rise to 7.5'C. To follow conditions in the Navesink River, we held water temperature at 7.5°C until 10 April, then raised temperature ~0.5°C per day to 15°C on 14 April. This temperature was main- tained until the end of observations on 13 May 1997. Finely chopped pieces of frozen surf clam iSpisiila solidissima} were made continuously available as food for the flounders. The food was renewed at least once per week during the early part of the period when feeding was light, and increased with inges- tion rate to near-daily additions by 1 April 1997. The food was placed on the bottom in an area approxi- mately 2 m in diameter within camera view so that all feeding bouts could be counted. Recording observations Two low-light-sensitive video-tape cameras were set at the end windows of the aquarium to make continuous recordings of court- ship and spawning events over a large portion of the tank from bottom to water surface. Each of these cameras viewed 53'7f (24.6 m-^) of the tank bottom with minimal overlap in coverage. Recordings were made from about 1530 h through 0500 h between 29 January and 18 April 1997, after preliminary taping and routine visual observation indicated that virtu- ally all activity was nocturnal. The validity of this approach was confirmed by cameras at the side win- dows set for 24-h recording throughout the study period. Twenty-four-hour recording from the end cameras was initiated 19 April 1997, concurrent with increasing daytime activity of the fish. Spawning events were divided into two classes. An event was considered "definite" when characteristic behavior patterns were clearly observed in the re- corded image. However, because most spawning oc- curred in darkness, some spawning events were not observed clearly. These likely but less certain spawn- ing events were scored "probable." Two forms of court- ship were recorded; these are described in the "Re- sults" section. Records were kept of the numbers of males and females involved in each courtship and spawning encounter. A feeding event was scored once, Stoner et al : Behavior of Pseudopleuronectes americanus during spawning, feeding, and locomotion 1001 regardless of feeding duration, when a fish entered the food distribution area and made obvious feeding motions to- ward the clam pieces. Analysis of video tape from the two cam- eras for four dates during midspawning season revealed that there were no significant differences in daily counts of spawning, courtship, and feed- ing as recorded by the two cam- eras (paired-sample t tests, P>0.30); therefore, the second camera was used only as a backup for recorder failure (e.g. on four dates). Two other low-light cameras were set at side windows to provide data on locomotory ac- tivity and swimming speeds. Each viewed a 3.6-m wide sec- tion of the opposite tank wall and about 4.6 m- of bottom, and recordings were made for 5 min at the beginning of each hour. An index of locomotory ac- tivity was provided by the number of fish passing a central line in view of the camera. Swimming speeds (cm/sec) were determined by timing the passage of fish between two points 142 cm apart on the tank wall. Whenever possible, the gender offish passing through this view was noted. The two cameras pro- vided essentially identical information, and record- ings from only one were analyzed for each day. Observations were also made on the relative abun- dance of winter flounder larvae in the aquarium. Flounder larvae were almost constantly present in the aquarium after mid-February, and an index of their abundance was recorded each day. Because the larvae demonstrated diurnal vertical migration, ob- servations were made at approximately 1200 h and general abundance was scored from zero to four. A score of one was recorded when only a few larvae were observed in any field of view and a score of four was given when thousands of larvae were present at all of the tank windows. Field collections To monitor the location, reproductive condition, and sex ratio of winter flounder in a wild population, weekly daytime otter-trawl surveys were conducted from 7 February to 24 April 1997 in the Navesink Jliver estuary. New Jersey (Fig. 1), which is a known spawning area for the species ( Scarlett' ). Eight fixed stations were established approximately 1 km apart Figure 1 The Navesink River estuary. New Jersey, showing the stations where collections were made for winter flounder during winter and spring 1997. throughout the system. Two 5-min tows (at 4.4 ±0.9 km/h) were made at each station with a 5-m semi- balloon otter trawl with 3.5-cm wings and body and a 6-mm codend liner. Abundance of winter flounder was reported in catch per unit of effort standardized as number of individuals per 100 m of tow distance, which was estimated with a global positioning sys- tem (GPS). Winter flounder collected in the trawl surveys were measured to the nearest millimeter of total length, blotted dry, and weighed to the nearest gram. Gen- der was determined by inspecting the gonads, and each fish was classified as immature, ripe, or spent. The gonads were weighed to the nearest gram, and gonadosomatic indices (GSI ) were calculated accord- ing to LeCren (1951) as GSI = 100 [gonad weight/total body weight). The stomach of each winter flounder was removed, preserved in 10% formalin solution, and later trans- ferred to 70'7f ethanol. Diet was analyzed as in Stehlik ( 1993), and fullness of the foregut was estimated vi- sually on a scale from 0% to 100%. Prey items were identified to the lowest possible taxon and the pro- portion of the total volume of stomach contents con- ' Scarlett, P. G. 1991. Temporal and spatial distribution of win- ter flounder ^ Pseudopleuronectes americanus) spawning in the Navesink and Shrewsbury Rivers, New Jersey. Dep. of Envi- ronmental Protection. Div. Fish. Game and Wildlife, Bureau of Marine Fisheries, Nacote Creek Marine Fisheries Laboratory, Star Route, Ahseron, New Jersey, 12 p. Unpubl. report. 1002 Fishery Bulletin 97(4), 1999 tributed by each prey type was estimated visually (Williams, 1981). Fullness was averaged for fish grouped by date, gender, and stage of reproductive development. Fish examined were 63-460 mm TL. Bottom water temperature, salinity, and dissolved oxygen were measured weekly at each station with YSI (Yellow Springs Instruments) oxygen and tem- perature-conductivity meters. HydroLab DataSonde II instruments were deployed near the bottom on piers extending from the south shore of the Navesink River near stations 1 and 6 to record temperature, salinity, and dissolved oxygen each hour. Results Laboratory observations Behavior Two forms of simple courtship behavior were observed in the research aquarium: "following" and "avoidance." "Following" was defined as oriented locomotion of one or more fish behind another, typi- cally on the sediment surface. A following could be relatively slow or fast (approx. 20-50 cm/sec), con- tinued for 5-20 sec, and sometimes resembled chas- ing. Turns by the lead fish were followed by the fish behind, and, in most cases, the distance between fish decreased but contact was never made. In all cases where genders of the fish could be determined the lead was female. One or more "followers" were al- ways male. "Avoidance" was characterized by brief contact (<1 s) between a "follower" and the female being followed, then by rapid acceleration and eva- sive turning by the female. Typically, the female swam 1-2 m off the bottom. In at least 509^ of such encounters additional males converged rapidly on the female, eliciting strenuous evasive swimming by the female. Our observations of spawning in winter fiounder were relatively similar to those of Breder ( 1922), ex- cept that females never initiated spawning. Spawn- ing occurred after initial contact by a male as de- scribed above. Instead of fleeing, the female remained near the sediment surface (<1.0 m above it) and the pair immediately made several very rapid circles with diameters typically less than two body lengths. Spawning events could be detected even in the far reaches of the tank during darkness because the fe- male nearly always rotated momentarily to the ver- tical position so that the white underside of her body was clearly visible on the outside of the circles. Par- ticipation by multiple males in the spawning event was common, and in approximately 107r of spawn- ing events, one to several males converged on the spawning location immediately after the female had departed, making rapid undulating motions on the sediment surface. Despite convergence of males on spawning females and spawning locations, agonistic behavior among males was never observed. In less than 29c of observed spawning events, a second spawning by the same female occurred a short dis- tance from the first encounter. In most cases the fe- male avoided further contact. Spawning Spawning began on 8 February 1997, about 2 weeks after the fish were introduced into the research aquarium. One or more spawnings oc- curred in camera view every day, except 14 Febru- ary, until 1 April (Fig. 2A); maxima were evident in mid-February and mid-March. The last spawnings were observed on 8 April. In total, 211 spawning events were recorded, 34% of which were definite. In analyses that follow, we considered the "spawning season" in the aquarium to be the period between 8 February and 8 April, 1997. The "postspawning sea- son" was the remainder of the observational period, 9 April through 13 May 1997. Spawning began near or just before sunset ( 1700- 1800 h) and reached maximum frequency around 2100 h (Fig. 3A). Seventy-five percent of all observed events occurred before midnight. Although some spawning occurred as late as 0430 h, only 10% oc- curred after 0200 h, when the fish became relatively inactive. When the total number of spawning events dur- ing the observation period was extrapolated to the total dimensions of the aquarium and divided by the ten females in the system, we estimated that the average female fiounder spawned 40 times during the reproductive season. It is unknown whether in- dividual fish spawned over this extended observa- tion period. However, given that the mean total num- ber of spawns per night was just six, individuals prob- ably spawned over a period of at least one week. Two facts suggest that a subset of the females may have been completely spent during the first three weeks of the spawning period: the nightly spawning fre- quency was bimodal, with a minimum on 2 March (Fig. 2A), and female feeding increased rapidly after approximately 3 March (Fig. 4). None of the observed spawning events involved more than one female flounder, but multiple males were engaged in the majority of spawnings. In 151 events with records for the number of males, only 22.5% were pair spawnings, 11.3% had two males, 3.3% had three males, and the majority involved four to six males (62.9%). Given these numbers, a conser- vative estimate for the total number of male spawnings within camera view is 780. By extrapolating to total tank dimensions and dividing among the ten males Stoner et al,: Behavior of Pseudop/euronectes amencanus during spawning, feeding, and locomotion 1003 12 1 a > (D J? c _ li II II C ^ XI ni (U M»t8 mg/L) at the two sites where measurements were made continuously. Population structure, reproduction, and migration Numbers of winter flounder in the Navesink River 1006 Fishery Bulletin 97(4), 1999 m Following ND . ND ND "ftrii iMiiininiii'niMnnni < < < < < ^- in o) (*5 3UU - 250- 200- B Avoidance 150- 100- SO- L^fcirf^ 0 ■ ND ill y^^^^ ^t^BKU^^m ^ ^ ■ NO ND ND ra 0) 0) "? H- 4- m (N 0.0.5. Source of variation df Sum of squares F ratio Date 9 0.002 3.27 '* Station 7 0.004 7.25 *** Gender 1 0.0003 4.62 * Date ■ Station 63 0.009 2.10 *** Date ■ Gender 9 0.0005 0.84 ■"■ Station - Gender 7 0.004 7.22 *** Date y Station ■ Gender 63 0.003 0.59 '"^ Error 160 0,011 lively Male winter flounder formed two size groups, small fish ( <20 cm TL ) assumed to be age- 1 fish and larger individuals (20-35 cm TL). Through the end of Feb- ruary all males larger than 20 cm were reproduc- tstuai ball (rally a(e itth Stoner et al.: Behavior of Pseudop/euronectes americanus during spawning, feeding, and locomotion 1007 1H A 12 10 H 6 i 4 2 0 Spawning period Female 111 I I 0 2 -1 — T — I — I — I — I — I — I — I — I — r" 6 8 10 12 14 16 18 20 22 14- C 12 - 10 8 6 H 4 2 0 Male I I I I I I I 0 2 4 6 T — I — 1 — I — I — r — I — I — 1 — r 8 10 12 14 16 18 20 22 14 12 10 8 6- 4- 2 0 B Postspawning period Female 6 8 10 12 14 16 18 20 22 16 18 20 22 Figure 6 Hourly frequency of feeding during spawning and postspawning periods for (A, B) female and {C, D) male winter flounder in the research aquarium. Values represent mean numbers of observed events per hour ±1 standard error The dashed vertical lines show the times of artificial sunrise and sunset in the aquarium at the midpoint of the spawning season. Photoperiod was maintained on a natural cycle for the latitude of the laboratory. tively ripe, as were 40% of males between 10 and 15 cm. A steady decline in male GSI began in March (Fig. 9B), but ripe males (mean GSI=7.3, SD=4.4) were col- lected later than ripe females, through 11 April. Although the number offish collected at any one station was not large, the Station x Gender interac- tion in the ANOVA indicated that male and female flounder had different distributional patterns in the estuary (Table 1). Males were more abundant than females at stations in the lower estuary (stations 7 and 8), whereas females were more abundant in the middle and upper reaches (stations 1-6) (Fig. 10). These gender-specific distribution patterns were gen- erally consistent, as indicated by the nonsignificant Date X Station x Gender interaction term in the ANOVA (Table 1). Ripe males were collected through- out the estuary in proportions ranging from 14% of individuals at station 6, to 100% at stations 1 and 2. In contrast, all ripe females were collected at sta- tions 1-4, with the exception of one partially spent female collected at station 7. The length distribution of female winter flounder at most of the sampling stations showed two modes (Fig. 11). Large females (mean length=29.7 cm, SD=0.6) were collected at all eight stations but were most abundant in the middle reach (stations 3-6). Small females (<20 cm TL), all of which were repro- ductively immature, were most abundant in the up- per estuary, with approximately equal numbers of small fish at stations 1-4. Size groups were less ob- vious in male winter flounder ( Fig. 11), but it is clear that the mean size of male fish was larger at sta- tions in the lower river (stations 5-8) than in the upper river, and small fish were most abundant at stations 1-3. From 28 March through the end of the sampling period, large numbers of flounders were collected only at station 6. Virtually all the fish in the lower river during this late part of the collecting season were spent. Feeding Stomach fullness in winter flounder var- ied with date, gender, and age (Fig. 12). Highest full- 1008 Fishery Bulletin 97(4), 1999 Spawning period Postspawning period Female r j 11 Figure 7 Locomotory activity in adult winter flounder in the research aquarium shown on an hourly basis for (A, B) females and (C, D) males during the spawning and postspawning periods. See text for the method used in generating this index of activity. The dashed vertical lines show the times of artificial sunrise and sunset in the aquarium at the midpoint of the spawning season. Photoperiod was maintained on a natural cycle for the latitude of the laboratory. ness indices (59%-84%) occurred in immature fish, most containing substantial amounts of food. Aver- age fullness in females was also high throughout the survey period (60'^-80%), but variation was large and associated with reproductive state. Ripe females had an average fullness of just 24*^, whereas spent females had an average fullness of 72%. Variation among males was even higher, with an obvious in- crease in fullness from about 1% early in the survey to a range of 70'/(-90'7r during the last three collec- tion dates (Fig. 12). Ripe males had average stom- ach fullness of only 17%, whereas average fullness of spent males was 69%. Therefore, the rapid increase in stomach fullness of males in late March was asso- ciated with rapidly declining numbers of ripe males. Stomach fullness was weakly correlated with GSI in both females (/•=-0.439) and males (r=-0.582) be- cause fullness in fish with a low GSI spanned the entire range from 0%-100%. Siphons of soft clam (Mya arenaria ) were the most abundant prey items taken by winter fiounder, on the basis of both percentage volume and frequency of occurrence. Of the 257 flounders that contained prey, 150 contained siphons with diameters of 3-13 mm. However, diet varied with collection site (Fig. 13). Siphons of M. arenaria were most abundant in fish collected in the middle reach of the estuary (sta- tions 3-6), whereas another clam, Macoma halthica, was most abundant in fish from the upper reach (sta- tion 1 ). Ampeliscid amphipods, mysids, shrimps, and other crustaceans were the most important food items in the lower estuary. h\ litiiii K(|) Hi Stoner et al : Behavior of Pseudopleuronectes americanus during spawning, feeding, and locomotion 1009 E 12 n 10 8 - 25-Jan 14-Feb 1^1 B 1.2 6-Mar 26-Mar 15-Apr 5-May o ^ 0.8 o 0.6 - d z 0.4 - 0.2 25-Jan 14-Feb 6-Mar 26-Mar 1 5-Apr 5-May Date Figure 8 (A) Water temperature and (B) abundance of winter flounder in the Navesink River estuary during winter and spring 1997. Temperature values are means and ranges of temperature at the eight sampling stations. Values for flounder abundance are mean catch (±1 standard error) per 100-m distance over the bottom sampled by a 5-m otter trawl. Discussion Spawning behavior Direct observation of spawning behavior has been documented for only a few flatfish, but spawning modes appear to vary substantially among species. Midwater pair-spawning has been observed in pla- ice (P. platessus; Forster, 1953), and in Dover sole iMicrostomus pacificus; BajTies et al., 1994), but win- ter flounder spawned near the bottom and pairwise spawning (22%) was less common than spawnings involving multiple males (78%). Also, winter floun- der spawning may be less elaborate or ritualized than in plaice or Dover sole or Caribbean bothids (Konstan- tinou and Shen, 1995 1. Although this could be an arti- fact of confinement, our research aquarium, with a bottom surface area of 46 m- and 3 m depth, was much larger than tanks where pair spawning has been observed in other species and where the den- sity and size offish were lower. Our long-term observation of a single spawning population in the aquarium allowed us to record sev- eral different forms of reproductive behavior not re- ported earlier (Breder, 1922). Most notably, we were able to observe and record individual encounters 1010 Fishery Bulletin 97(4), 1999 ■ 45 -| A • 40 - , • Female 35- 30 - : 1 • • ripe 25 - 20 - 15 - • • 0 spent index {%) 3 Ol o ^ M s 1 1 § 1 ° '^ ° e S 1 § o 1 25- o 25 T a s o Jan 14-Feb 6-Mar 26-Mar 15-Apr 5-May B 20- 15 - 10 - • Male • • • • • • ripe o spent . 1 . • 5 - • : 1 : : ! • ° n - • = • 6 i a e 5 25- Jan 1 1 1 r — 1 14-Feb 6-Mar 26-Mar 15-Apr 5-May Date Figure 9 Gonadosomatic index ( GSI ) for winter flounder (A) females and (B) males shown as a function of collection date. The index was calculated according to LeCren (1951) as GSI = lOOigonad weight/total body weight). Observational conclu- sions about the reproductive state of individual fish are shown by dots for ripe fish and circles for spent fish. Note the different ,v-axes. veri toil !ti! IS II byi mo! acti rem cuei tliei visi UtCi tof between males and females. Unlike Breder, and in thousands of observed encounters, we never observed females initiating spawning activities. Breder ac- knowledged that his relatively small tanks may have altered normal behavioral patterns and was correct in predicting that males would be more active than females in a more open environment. In fact, males attempted to initiate spawning in the laboratory long before and after females were receptive. A similar phenomenon most likely occurs in the field, where males are ripe and capable of spawning for a longer period of time than females. Similarly, Rijnsdorp ( 1989) has observed that male plaice are in spawn- ing condition for a period twice as long as that for females. As in our observations with winter floun- der, Baynes et al. (1994) observed that males of Microstomus were much more active than females, and male attempts to initiate spawning were often rejected. Male winter flounder were able to detect the pres- ence of a spawning female from distances of at least 5-10 m in the research aquarium and frequently con- pen mm weri live 'lers evei avei Dec( terl ors 1981 Si rang orjt ier rli)1 (152 ami time Stoner et al : Behavior of Pseudopleuronectes americanus during spawning, feeding, and locomotion 101 1 verged suddenly on a spawning pair or group. A variety of sensory mechanisms could play a role in detecting opportunities for egg fertilization. Male fish may respond to the obvious visual signal of light reflected from the white underside of spawning fish. It is also possible that hydroacoustic cues were emitted from vigorous bouts of spawn- ing that occurred just above the bottom. It is less likely that spawning was detected by an olfactory cue because there was al- most instantaneous response to spawning activity and there was little horizontal cur- rent in the aquarium. However, olfactory cues may be important in certain circum- stances. Occasionally, males converged on locations where spawning had occurred al- though the spawners had departed, and these spawning locations were sometimes visited by additional males for several min- utes after the initial spawning event. Fu- ture research will be needed to determine the precise mechanisms of spawning group formation and detection. In female winter flounder just one batch of oocytes matures per year (Burton and Idler, 1984); never- theless, spawning in the aquarium population of win- ter flounder had a duration of 60 days. It is clear that individual females spawn many times over a period of at least one week, and perhaps longer. Cir- cumstantial evidence indicates that individual males were capable of spawning over most of the reproduc- tive season in the research aquarium. Large num- bers of males were often engaged in single spawning events; the total number of spawnings yielded a high average spawning frequency by males that would necessitate several weeks of spawning by each; and feeding by males did not begin until most females stopped spawning. It is unknown whether male win- ter flounder produce more than one batch of sperm or simply release one batch over time (Bert et al., 1988). Spawning in the aquarium occurred over a broad range of times, but always during hours of darkness or just before. Nocturnal spawning in winter floun- der is probably a function of endogenous diurnal rhythms rather than light level because Breder (1922) noted that spawning occurred between 2200 and 0330 h even under artificial light. Spawning times vary with flatfish species and are probably species-specific. For example, plaice (Nichols, 1989) and other pleuronectids (Woodhead, 1966) are also known to spawn primarily at night, but Bothus ocellatus spawns only around the time of sunset (Konstantinou and Shen, 1995). 1 - 0.9- 0.8 - 0 7 - E 8 0.6- 1 0.5 ■ o o 04- Z 0.3 ■ 02 - 0 1 - A -^Female -^Male J}j \ / 12 3 4 5 6 7 8 station Figure 10 Mean catch by station number of male and female winter flounder per 100 m distance over the bottom in the Navesink River estuary. Values plotted are means for all dates combined. The role of multiple releases of eggs and sperm is poorly understood but may be important in the re- productive biology and recruitment of winter floun- der. Although release of gametes could not be re- corded or verified beyond the fact that larvae were produced in the aquarium, it is evident that numer- ous males spawned in response to one female. It is likely, therefore, that eggs released in a single spawn- ing event were often fertilized by more than one male. Given that the average female fish spawned 40 times during the reproductive season in the aquarium and that most of the males appeared to remain ripe throughout the observation period, it is likely that many, if not all, of the males could have contributed genetic material to the offspring of each female. Ge- netic diversity of embryos produced in the field could be high for similar reasons. Although male winter flounder appeared to be in a constant state of readiness for spawning, future experiments will be needed to determine whether frequency of spawning and fertilization success is uniform among individuals and how this might be influenced by size structure, sex ratio, or density of fish on the spawning ground. For example, small ( 10— 15 cm ) but reproductively ripe males were abundant in the Navesink River estuary. It is possible that these small individuals have a lower frequency of success than large males in inducing females to spawn, given the order-of-magnitude difference in body weight between 10 and 22 cm fish. However, Baynes et al. (1994) observed female Dover sole spawning with males smaller than themselves, and 1012 Fishery Bulletin 97(4), 1999 Female Male n = 14 at n t^ « a a -1 — ■ — 1 — I 0 5 » n =3 1 mims n o> « h. zr. IP. 91 Q K U2 n =16 O) n r«- 4 0) O (1 o (1) n i;^ F 3 Z 8 n =28 4 ^Ui ^ — ^ ^^ f^ f^^ fwj^p^l^ Ol (O T«- tN CM R a n It 4 • 0 n =18 m, — fB, — ^ n =9 O) r) r- « S a f; ; « n =5 4 P^ p!!f«I^ 1 W 1 1 1— o> P> 1^ « R S ft 5 !« n =9 5 — , — t^f™, — , — ,- n n ■^ n =41 6 m ffl W Pfw^ , , , , P 1 Ip p, 12 « a a ft 5 « n =38 i 6 l^fl p. p,y^^ f 1^ W oi n f^ 12 fi R a ft n =5 7 1 1 1 1 1 1 ^t- n???ir?HS . m rssi, , en n r^ »- ui en n ^^ ^- in oj rj n n ^ ^ n =7 n =20 7 , m , ^^ ^m» A^l^^^ o> n r^ oi (O n ■v ^ n =36 0 I W , r- 9 CT) n r^ Total length (cm) Figure 11 Length-frequency distributions for female and male winter flounder in the Navesink River estuary, plotted by field station. the same may occur with winter flounder. On the other hand, most fertihzation by small males may occur during group spawning. "Sneaking," whereby subordinate males approach and fertilize some of the eggs when a pair is spawning, is a reproductive behav- ior common in a variety of fishes (Chan and Ribbink, 1990; Oliviera and Almada, 1998 ). We observed no overt agonistic behavior in our long-term experiment, and there is no reason to believe that small males do not contribute to the reproductive population. Stoner et al : Behavior of Pseudopleuronectes americanus during spawning, feeding, and locomotion 1013 Spawning, migration, and habitat use Previous studies have shown that winter flounder in the mid-Atlantic region migrate to shallow inshore waters to spawn during the late win- ter (Perlmutter, 1947; Poole, 1966; Scarlett and Allen, 1992), and it is known that the Navesink-Shrews- bury River estuarine system is an important spawning ground ( Phelan, 1992; ScarlettM. Phelan's study and our investigation provide direct evi- dence that winter flounder immi- grate to the Navesink estuary be- tween early February and March. Flounder abundance was lowest when water temperature was lowest (2.2'^C in mid-February) and in- creased rapidly with subsequent tem- perature rise. In fact, high and rela- tively constant catches of winter flounder occurred in the Navesink estuary at temperatures of 6-ll°C. This is consistent with earlier obser- vations that adult winter flounder emigrated from bays in southern New Jersey when water temperature declined below 3.5"C and returned when temperature rose to 6°C (Danila, 1978). Most adult males and females had ripe gonads upon entry to the system in late winter, and most were spent by the end of April. This pat- tern of seasonality is consistent with previous findings (Pearcy, 1962; Scarlett and Allen, 1992). It is likely that the most important spawning area for winter flounder in the Navesink River was in the middle reach. Reproductively ripe male win- ter flounder were found throughout the estuary; however, highest num- bers of ripe females were found west of station 5, and only one nearly spent female was collected farther down the estuary. Scarlett and Allen ( 1992 ) found that the middle reach of the Manasquan River estuary. New Jersey, was a major spawning ground for winter flounder, as were the upper reaches of the Mystic River estuary, Connecticut (Pearcy, 1962 ), and Narragansett Bay, Rhode Island ( Powell- ). 100 - 80 - S 60- (D Q. -A 1 • 1 ^ 1 -•-Females _2 1 1 -o- Males 1 40- 1 - A- Immature a5 20 ■ a / / / / 0^ ^ 14-Feb 6-Mar 26-Mar 15-Apr 5-May Date Figure 12 Temporal variation in percent stomach fullness of winter flounder in the Navesink River estuary. Fullness indices are shown by adult gender and for immature fish. kW?-Ampefisca sp CRU=Crustacea HYD=Hydrozoa MAC=/Macoma sp MIS=Miscellaneous MOL=Mollusca MYS=Mysiclacea MYA= Uya arenana POL=Polychaeta Figure 13 Composition of diet by percentage volume in winter flounder collected at eight stations along the length of the Navesink River estuary during winter and spring 1997. - Powell, J. C. 1989. Winter flounder tagging study. 1986- ^ 1988, with comments on movements. Rhode Island Division of Fish, Wildlife, and Estuarine Research. P.O. Box 218, West Kingston, Rhode Island 02892. Unpubl. report. Several different physical and chemical factors could affect the location of spawning habitats. Tem- perature may have influenced the seasonal migra- tion to the Navesink estuary; however, there was little variation in temperature along the axis of the estuary, and other factors probably influenced exact spawning location. McCracken (1963) found that adult winter flounder died in 8 ppt salinity; therefore, upstream migration is probably limited by this variable. Neither males nor females were abundant in locations where salinity was frequently near or below 10 ppt. 1014 Fishery Bulletin 97(4), 1999 The location of spawning in an estuary may also be related to temperature and salinity requirements of the demersal eggs. Scarlett and Allen ( 1992 ) col- lected eggs at salinities of 14-32 ppt and tempera- tures of 0.9-10°C in the Manasquan estuary, about 30 km south of our study site. Rogers (1976) found that the highest proportion of viable hatches occurred at 3°C over a salinity range of 15-35 ppt. At tem- peratures over 3°C, the optimal range was 15-25 ppt, which closely matches temperature and salinity ranges in the middle reach of the Navesink River Other factors that may influence the choice of spawning habitat by winter flounder are hydrody- namic properties and associated sediment character- istics. Strong flood tidal currents that characterize the lower Navesink River attenuate rapidly in the middle reach (Chant'^). This attenuation is associ- ated with deepening of the estuary and general re- duction of sand bars, decrease in the mean grain size of sediments, and increase in sediment organic con- tent (Stoner, unpubl. data). It is unknown whether spawning winter flounder respond to any of these habitat characteristics; however, the depositional qualities of the upper estuary and flood-dominated circulation may aid in the retention of winter floun- der eggs and the evolution of habitat choice in spawn- ers. Crawford and Carey (1985) noted that winter flounder spawned in two sites in an estuarine lagoon in Rhode Island where hydrodynamic features re- tained larvae. It is plausible that habitat choice, cued by either fine-grained sediments or low cuiTent veloci- ties, could be associated with such retention features. Seasonal variation in feeding activity The general conclusion has been that winter floun- der are opportunistic feeders, consuming polychaetes, small crustaceans, molluscs, and other prey accord- ing to their abundance (Pearcy, 1962; Richards, 1963 Klein-MacPhee, 1978; Franz and Tancredi, 1992 Martell and McClelland, 1994; Steimle et al., 1994 Carlson et al., 1997 ). Winter flounder collected in the Navesink River during this investigation consumed a mixture of invertebrate prey, but the vast majority ofthe diet was siphonsof the soft clam Mya arenaria. The abundance of spent and feeding flounder in the lower estuary may. in fact, be associated with the observed abundance of this preferred prey in that location. Cessation of feeding by adult winter flounder dur- ing the winter has been noted previously ( Tyler, 1972; ^ Chant, R.J. 1998. Instituteof Marine and Coastal Sciences, Rutgers University, P.O. Box 231, New Brunswick, New Jersey 0890.3. Unpubl data. Martell and McClelland, 1994), but we found that feeding was influenced strongly by gender and re- productive state. Both in the field and in the aquarium, female flounder appeared to resume feed- ing immediately after they completed spawning for the season. Early feeding probably provides an ad- vantage for female fish because ovarian development for the next spawning season begins immediately after spawning (Burton and Idler, 1984). In the field, males did not begin feeding until the end of March, and feeding in the aquarium was light until after the last spawning events. We conclude that male behavior is dedicated to fertilizing the maximum number of eggs, at the expense of feeding and late winter growth. There is now substantial evidence for seasonal variation in daily activity and feeding rhythms in winter flounder During the summer, winter flounder are primarily diurnal in general activity (011a et al., 1969) and feeding (MacDonald and Waiwood, 1987). Bharadwaj (1988) found that adults fed throughout the day and night in November, speculating that this was related to maturation of gonads for the winter spawning season. In our aquarium observations, adult winter flounder were almost entirely noctur- nal throughout the spawning season, remaining bur- ied during the day. Daytime activity began again when spawning was completed and increased throughout the remainder ofthe postspawning period. Primarily daytime activity during the nonspawning season could be associated with visual predation modes (Olla et al., 1969), but it remains to be determined how mates are located in darkness and why spawn- ing occurs at night. Conclusions This study provides the first detailed observations on winter flounder spawning since Breder's (1922) original description over 75 years ago, and the com- bination of laboratory and field work provides new insight into the biology and ecology of spawners. The behavior of male winter flounder appears adapted to maximize encounters with reproductive females and numbers of eggs fertilized. From our field collec- tions, it is apparent that one-year-old male winter flounder ( 10 cm TL) are capable of spawning, whereas female spawners all appeared to be at least two years old. Males also became reproductively ripe earlier in the season than did females and remained ripe to a later date than most females. Males were more ac- tive than females, had higher swimming speeds, and initiated all spawning events. The males fed very little during the reproductive season and appeared i Stoner et al.: Behavior of Pseudopleuronectes amencanus during spawning, feeding, and locomotion 1015 to sacrifice winter growth to maintain a high level of reproductive behavior. Spawning frequency was high in both males and females. Individual females were capable of spawn- ing as many as 40 times over a period of at least one week, and males spawned throughout the season (at a rate at least three times the female rate) and per- haps several times per day. Clearly, individual records of spawning behavior would be useful in un- derstanding the reproductive role of one-year-old males and in determining the duration of spawning in individual females. Given the frequency of spawn- ing in both male and female winter flounder and the fact that several males may release sperm in re- sponse to egg-laying by one female, we conclude that the genetic diversity of offspring produced by a fe- male in one year is probably very high. Although our data provide new insights into the natural history and reproductive behavior of winter flounder, several questions important to sound man- agement of spawning stocks remain to be answered. For example, we do not know if reproductive popula- tions are scattered in estuarine and coastal waters or are concentrated in spawning aggregations on specific spawning grounds. We saw a concentration of ripe fish in the middle reach of the Navesink River estuary, but ripe fish were also present in a variety of habitats in adjacent Raritan Bay during the win- ters of 1997 and 1998 (Stoner and coauthors, per- sonal obs.), and it is possible that winter flounder spawn in more than one location over the spawning season. It is also important to learn whether or not spawning behavior and reproductive output from a spawning ground are density-dependent. Acknowledgments We wish to thank D. McMillan and the staff of the Coastal Ecology Branch for help in collecting adult winter flounder for the experiments, J. Vitaliano for assistance with video systems, and T. Kalmar for help in analyzing video tapes. We are also grateful to the maintenance staff at the Howard Laboratory, who provided dependable seawater and lighting systems for the long duration of our aquarium study. B. OUa, A. Studholme, and anonymous reviewers provided helpful criticism of the manuscript. Literature cited Able, K. W., and M. P. Fahay. 1998. The first year in the Hfe of estuarine fishes in the Middle Atlantic Bight. Rutgers Univ. Press, New Brunswick, NJ, 342 p.. Baynes, S. M., B. R. Howell, T. W. Beard, and J. D. Hallam. 1994. A description of spawning behaviour of captive Do- ver sole, Solea solea (LJ. Neth. J. Sea Res. 32:271-275. Bert, A., D. L. Kramer, K. Nakatsuru, and C. Spry. 1988. The tempo of reproduction in Hyphessobrycon piilchriplnnis (Characidae), with a discussion on the biol- ogy of 'multiple spawning' in fishes. Environ. Biol. Fishes 22;15-27. Bharadwaj, A. S. 1988. The feeding ecology of the winter flounder, Pseudo- pleuronectes cimencanus (Walbaum). in Narragansett Bay, Rhode Island. M.S. thesis, Univ Rhode Island. Kingston, RI, 129 p. Breder, C. M. 1922. Description of the spawning habits of Pseudopleuro- nectes americanus in captivity. Copeia 102:3-4. Burton, M. P., and D. R. Idler. 1984. The reproductive cycle in winterflounder,Pseudop/euro- nectes americanus (Walbaum). Can. J. Zool. 62:2,563- 2.567. Carlson, J. K., T. A. Randall, and M. E. Mroczka. 1997. Feeding habits of winter flounder iPleuronectes americanus) in a habitat exposed to anthropogenic disturbance. J. Northw. Atl. Fish. Sci. 21:65-73. Chan, T.-Y.. and A. J. Ribbink. 1990. Alternative reproductive behaviour in fishes, with particular reference to Lepomis macrochira and Pseudo- crenilabrus philander. Environ. Biol. Fishes 28:249-256. Crawford, R. E., and C. G. Carey. 1985. Retention of winter flounder larvae within a Rhode Island salt pond. Estuaries 8:217-227. Danila, D. J. 1978. Age, growth, and other aspects of the life history of the winter flounder, Pseudopleuronectes americanus ( Walbaum ), in southern New Jersey. M.S. thesis, Rutgers Univ., New Brunswick, NJ, 79 p. Forster, G. R. 1953. The spawning behaviour of plaice. J. Mar. Biol. Assoc. U.K. 32:319. Franz, D. R., and J. T. Tancredi. 1992. Secondary production of the amphipod Ampelisca abdita Mills and its importance in the diet of juvenile win- ter flounder [Pleuronectes americanus) in Jamaica Bay, New York. Estuaries 15:193-203. Howe, A. B., and P. G. Coates. 1975. Winter flounder movements, growth, and mortality off Massachusetts. Trans. Am. Fish. Soc. 104:13-29. Kennedy, V. S., and D. H. Steele. 1971. The winter {\ounderi Pseudopleuronectes americanus) in Long Pond, Conception Bay Newfoundland. J. Fish. Res. Board Can. 28:1.153-1,165. Klein-MacPhee, G. 1978. Synopsis of biological data for the winter flounder. Pseudopleuronectes americanus (Walbaum). U.S. Dep. Commer, NOAA Tech. Rep. NMFS Circ. 414. 43 p. Konstantinou, H., and D. C. Shen. 1995. The social and reproductive behavior of the eyed flounder. Bothus ocellatus. with notes on the spawning of Bothus lunatus and Bothus ellipticus. Environ. Biol. Fishes 44:311-324. LeCren, E. D. 1951. The length-weight relationship and seasonal cycle in gonad weight and condition in the perch (Perca fluciattlis). J, Anini. Ecol. 20:201-219. MacDonald, J. S., and K. G. Waiwood. 1987. Feeding chronology and daily ration calculations for 1016 Fishery Bulletin 97(4), 1999 winter flounder (Pseudopleuronectes americanus). Ameri- can plaice iHippoglossoides platessoides). and ocean pout (Macrozoarces americanus) in Passamaquoddy Bay. New Brunswick. Can. J. Zool. 65:499-503. Martell, D. J., and G. McClelland. 1994. Diets of synipatric flatfishes, Hippoglossoides plates- soides. Pleuronectes ferrugtneus, Pleuronectes americanus, from Sable Island Bank, Canada. J. Fish Biol. 44:821- 848. McCracken, F. D. 1963. Seasonal movements of the winter flounder, Pseudo- pleuronectes americanus (Walbaum) on the Atlantic coast. .J. Fish. Res. Board Can. 20:551-586 Nelson, S. A., J. E. Miller, D. Rusanowsky, R. A. Greig, G. R. Sennefelder, R. Mercaldo-Allen, C. Kuropat, E. Gould, F. P. Thurberg, and A. Calabrese. 1991. Comparative reproductive success of winter flounder in Long Island Sound: a three-year study ( biology, biochem- istry and chemistry). Estuaries 14:.318-331. Nichols, J. H. 1989. The diurnal rhythm in spawning of plaice (Pleuro- nectes platessa L.I in the southern North Sea. J. Cons. Int. Explor Mer 45:277-283. Oliviera, R. F., and V. C. Almada. 1998. Mating tactics and male-male courtship in the lek- breeding cichlid Oreochromis mossambicus. J. Fish Biol. 52:1,11,5-1,129. Olla, B. L., R. Wicklund, and S. Wilk. 1969. Behavior of winter flounder in natural habitat. Trans. Am. Fish. Soc. 98:717-720. Pearcy, W. G. 1962. Ecology of an estuarine population of winter floun- der, Pseudopleuronectes americanus (Walbaum). Bull. Bingham Oceanogr. Collect. Yale Univ. 18:5-78. Perlmutter, A. 1947. The blackback flounder and its fishery in New En- gland and New York, Bull. Bingham Oceanogr. Collect. Yale Univ. 11(2), 92 p. Phelan, B. A. 1992. Winter flounder movements in the inner New York Bight. Trans. Am. Fish. Soc. 121:777-784. Pierce, D. E., and A. B. Howe. 1977. A further study on winter flounder group identifica- tion off Massachusetts. Trans. Am. Fish. Soc. 106:131- 139. Poole, J. C. 1966. Growth and age of winter flounder in four bays of Long Island. N.Y Fish Game J. 13:206-220. Richards, S. W. 1963. The demersal fish population of Long Island Sound. Bull. Bingham Oceanogr. Collect. Yale Univ,18:l-101. Rijnsdorp, A. D. 1989. Maturation of male and female North Sea plaice (Pleuronectes platessa L.). J. Cons. Int. Explor. Mer. 46:3,5-51. Rogers, C. A. 1976. Effects of temperature and salinity on the survival of winter flounder embryos. Fish. Bull. 74:52-58. Saila, S. B. 1961. A study of winter flounder movements. Limnol. Oceanogr 6:292-298. Scarlett, P. G., and R. L. Allen. 1992. Temporal and spatial distribution of winter flounder (Pleuronectes americanus) spawning in Manasquan River, New .Jersey Bull. N.J. Acad. Sci. 37(1):13-17. Scott, W. B., and M. G. Scott. 1988. Atlantic fishes of Canada. Can. J. Fish. Aquat. Sci. 219, 731 p. Stehlik, L. L. 1993. Diets of the brachyuran crabs Cancer irroratus, C. borealis, and Ovalipes ocellatus in the New York Bight. J. Crustacean Biol. 13:723-735. Steimle, F. W.. D. Jeffress, S. A. Fromm, R. N. Reid, J. J. Vitaliano, and A. B. Frame. 1994. Predator-prey relationships of winter flounder, Pleuronectes americanus, in the New York Bight apex. Fish. Bull. 92:608-619. Tyler, A. V. 1972. Food resource division among northern, marine, de- mersal fishes. J. Fish. Res. Board Can. 29:997-1,003. Williams, M. J. 1981. Methods for analysis of natural diet in portunid crabs (Crustacea, Decapoda, Portunidae). J. Exp. Mar Biol. Ecol. 52:103-113. Witherell, D. B., and J. Burnett. 1993. Growth and maturation of winter flounder Pleuronectes americanus. in Massachusetts. Fish. Bull. 91:816-820. Woodhead, P. M. J. 1966. The behaviour offish in relation to light in the sea. Ann. Rev. Oceanogr. Mar Biol. 4:337-403. 1017 Abstract. -The humpback whale tMcgaptera nocaeangliae) is a cosmo- pohtan species whose stocks were dras- tically decreased by conimercia! whal- ing practices prior to 1967. The North Pacific population was estimated to be between 15.000 and 20.000 animals before the practice of whaling. At the time of the commencement of its inter- national protection in 1967, this popu- lation may have been reduced to fewer than 1000 individuals. The Pacific coast of Mexico and the Revillagigedo Archi- pelago constitute one of the main breed- ing and calving areas for North Pacific humpback whales. The objective of this paper is to present an estimation of abundance of humpback whales in this region based on photographic identifi- cation of individual animals. Estimates of population size were obtained by us- ing mark and recapture models for both closed and open populations, with each year representing a capture occasion. A total of 1184 humpback whales were identified in Mexican waters between 1986 and 1993. The best estimates of population size for the Mexican stocks were those provided by the modified Jolly-Seber method: 1813 OS^r CI: 918- 2505) for the coastal stock in 1992, and 914 (95'y CI: 590-1193) for the Revil- lagigedo stock in 1991. Population size of humpback whale, Megaptera novaeangliae, in waters off the Pacific coast of Mexico Jorge Urban R. Departamento de Biologia Marina Universidad Autonoma de Baia California Sur Ap Post 19-B La Paz, BC-S. 23081 Mexico E-mail lurbana'calafia uabcs mx Carlos Alvarez F. Mario Salinas Z. Laboratorio de Mamiferos Marines Universidad Nacional Autonoma de Mexico Ap, Post 70-572 Mexico, D F 04510 Mexico Jeff Jacobsen PO Box 4492 Areata, California 95521 Kenneth C. Balcomb III Center for Whale Research 1359 Smugglers Cove Road Fnday Harbor, Washington 98250 Armando Jaramillo L. Departamento de Biologia Marina Universidad Autonoma de Baia California Sur Ap. Post 19-B La Paz, B.C.S- 23081 Mexico Paloma Ladron de Guevara P. Anelio Aguayo L. Laboratorio de Mamiferos Marines Universidad Nacional Autonoma de Mexico Ap Post 70-572, Mexico, DP 04510 Mexico Manuscript accepted 15 January 1999. Fish. Bull. 97:1017-1024 ( 1999)'. Humpback whales, Megaptera no- vaeangliae, make seasonal migra- tions between low-latitude winter- ing areas used for mating and calv- ing and high-latitude feeding areas. The general distribution of feeding areas in the North Pacific covers coastal waters in the western North Pacific from northern Japan throughout the Bering Sea and in the eastern North Pacific as far south as southern California. Dur- ing the winter breeding season, these whales congregate in three geographically isolated tropical ar- eas: the Ryukyuan, Bonin, and Mariana Islands south of Japan; the islands of the Hawaiian Archi- pelago: and the Pacific coast of Mexico and the Revillagigedo Archi- pelago (Rice, 1974; Johnson and Wolman, 1984). 1018 Fishery Bulletin 97(4), 1999 Humpback whales have been hsted as endangered since severe reduction of all stocks worldwide by com- mercial exploitation (Rice, 1974; Gambel, 1976). The number of these whales were estimated to be between 15,000 and 20,000 animals before whaling depleted them during the first half of the 20th century; at the time of its international protection in 1967 this popu- lation may have been reduced to fewer than 1000 individuals (Rice, 1974, 1978). However, the reliabil- ity of these figures is unknown. The extent of the recovery in the North Pacific population over the last 25 years is debatable. Esti- mates of abundance with mark and recapture tech- niques based upon photo-identification data have been made for different areas of this ocean, but there has been much debate regarding the reliability of such estimates! Darling etal., 1983; Baker etal., 1986; Darling and Morowitz, 1986; Baker and Herman, 1987; Alvarez et al., 1990; Calambokidis et al., 1990; Cerchio, 1998; Calambokidis et al.^). There is cuiTently insuffi- cient evidence to assess whether a significant change in abundance has occurred since whaling ceased. Three main wintering aggregations of humpback whales are recognized off the Pacific coast of Mexico: the Baja California Peninsula; the mainland coast of Mexico (including, Isabel Island, Tres Marias Is- lands, and the mainland coast); and the Revillagigedo Archipelago (including Socorro, San Benedicto, Roca Partida, and Clarion Islands) (Rice, 1979; Urban y Aguayo, 1987). The comparison of photo-identified humpback whales in these wintering gi-ounds showed that there was a greater affinity between whales off Baja Cali- fornia and those off the mainland coast of Mexico than those off either Baja California or the main- land coast and those off the Revillagigedo Archipelago (Urban et al., 1989; Ladron de Guevara et al., 1993; Jaramillo, 1995) (Table 1). From these results, two different population units of humpback whales dur- ing winter in Mexican waters were previously pro- posed: the coastal Stock (including Baja California and mainland coast of Mexico), and the Revillagigedo stock (Alvarez et al., 1990; Urban et al.^) (Fig. 1). ' Calambokidis, J., G. H. Steiger, .J. Straley. T. J. Quinn II, L. M. Herman, S. Cerchio, D. Salden, M. Yagamuchi, F. Sato, J. Urban R., J. K. Jacobsen, O. von Ziegesar, K. C. Balcomb, C. M. Gabriele, M. E. Dahlheim, N. Higashi, S. Uchida, J. K. B. Ford, Y. Miyamura, P. Ladron de Guevara P., S. A. Mizroch, L. Shlender, K. Ra.smussen. 1977. Abundance and population structure of humpback whales in the North Pacific ba.sin. Final Kep. to Southwest Fisheries'Science Center, Natl. Mar Fish. Serv., NOAA, La JoUa, CA, 72 p. 2 Urban R., J., J. C. Salinas V., A. Guillen G., and E. Vazquez M 1997. La ballcna jorobada Mcgciptera novcwanfiluw en al Penin- sula de Baja California Sur. Mexico. Final Report to the Bio- diversity National Commission (CONABIOi. Contract Il():3.'>, 41 p. Table 1 Matches (above the diagonal) and interchange index' i be- low the diagonal) among the three main wintering aggi-ega- tions in the Mexican Pacific. Sample size in parentheses. Baja Mainland Revilla- California coast of gigedo Peninsula Mexico Archipelago (471) (383) (4501 Baja California Peninsula — 64 20 Mainland coast of Mexico 0.38 — 23 Revillagigedo Aj-chipelago 0.12 0.18 — ' Index of interchange. This index quantify the degree of interchange among two samples (among regions) that accounted for sample size: Index of interchange = (m^.^Kn, x njV x 1000, where/!, = whales identified (captured) in sample 1; n.y = whales identified in sample 2; and ni.,= captured whales from sample 1 recaptured in sample 2, (see Baker et al. 1985; Cerchio et al , 19981 This division of winter aggregations is also supported by mitochondrial DNA lineage analyses (Medrano- Gonzalez et al., 1994, 1995a, 1995b); however, be- cause the waters off California, Oregon and Wash- ington are the primary feeding destination of the whales observed in Baja California and the main- land coast, the migratory destination of the whales seen around the Revillagigedo Islands is still un- known (Urban et al. 1987; Urban et al.'^). Previous estimates of humpback whale abundance in Mexican waters are the following: 500-600 for Socorro Is. (Campos, 1987); 200-400 for Isabel Is. (Alvarez, 1987); 600-700 for the mainland coast (Alvarez et al., 1990); and 1200-1700 for all the Mexi- can Pacific (Urban et al., 1989). However, it is recog- nized that these estimates were based on limited data and on assumptions that generally were not tested. In our study we present an analysis of photographic data obtained during the winter breeding and calv- ing seasons from 1983 to 1993 in the different as- sembly areas of the Mexican Pacific. We use this analysis to calculate reliable independent estimates of abundance of humpback whales for the coastal and the Revillagigedo stocks. ' Urban R., J., A. Jaramillo L., A. Aguayo L., Paloma Ladron de Guevara P., M. Salinas Z., C. Alvarez F., L. Medrann G., J. Jacobsen, K. C. Balcomb, D. E. Claridge, J. Calambokidis, G. H. Steiger, J. M. Straley, O von Ziegesar, S. Mizroch, .M. Dahlheim, J. M. Waite, J. D. Darling, and C. S. Baker. 19xx. Migratory destination of the Mexican Pacific humpback whales. Universidad Autdnoma de Baja California Sur, Ap. Post. 19-B, La Paz, Baja California Sur, Mexico. Urban R. et al.: Megaptera novaeangliae in waters off the Pacific coast of Mexico 1019 ■24° Baja California Peninsula -20° Clari6n Is. 115° _] Revillagigedo Stock • San Benedicto Is. o Soconx) Is. 110° _J 105" _J Figure 1 Study area showing, by shading, the distribution of the Revillagigedo and coastal stocks. Materials and methods Individual identification Humpback whales were individually identified from photographs taken by different institutions from approximately December to March, along the main- land coast of Mexico (1983-92), from mid-January to March in Baja CaUfornia (1987-93), and at the Revillagigedo Archipelago (1986-92) (Table 2). Although photo-identification data have been available for the coastal stock since 1983 and for the Revillagigedo stock since 1985, efforts before 1986 were limited in time and space, and also lacked con- tinuity. Beginning in 1986, surveys were conducted off Socorro Island for the Revillagigedo stock, and near San Jose del Cabo or the mainland coast (or both) for the coastal stock. No data were obtained for either Revillagigedo or the mainland in 1993. The whales were identified individually through photographs of the black and white pigmentation pat- terns on the ventral surface of their flukes (see Katona and Whitehead, 1981 ). Photographs were obtained with 35-mm cameras equipped with 200—300 mm lenses. Although the film used varied by region and season, most photographs were taken with black and white Kodak Tmax 400 ISO pushed to 1600 ISO, ensuring shutter speeds as high as 1/1000 of a second. On the basis of focus, angle, and light conditions, all fluke photographs were judged to be either good, fair, or of poor quality; only photos in the first two categories were included in this study. Within these quality levels, the whales showed at least 50% of each fluke at a sufficiently vertical angle to allow the shape of the trailing edge to be distinguished. Calves were excluded from the analysis owing to their tendency to show changes in pigmentation patterns during the first year of life (Carlson et al., 1990). Selected pho- tographs were compared visually by at least three persons with experience in matching humpback whales flukes photographs from both Universidad Nacional Autonoma de Mexico and Universidad Autonoma de Baja California Sur. Abundance estimation Abundance estimates were obtained by using an eight-year period for the coastal stock, from 1986 to 1993, and a seven-year period for the Revillagigedo stock, from 1986 to 1992. Estimates of population size were obtained by us- ing mark and recapture models for both closed and open populations, with each year representing a cap- ture occasion. The time span of the study was seven years; consequently, although calves were not in- cluded in the analysis, it is inevitable that additions 1020 Fishery Bulletin 97(4), 1999 Table 2 Study periods and data sources for each of the regions UNAM = Universidad Nacional Autonoma de Mexico; UABCS = Universidad Autonoma de Baja California Sur; CWR = = Center for Whale Research; CS =Cousteau Society. Region Season Study period Source 1986 20 Dec 85-4 Mar 86 UNAM ^" 1987 25 Dec 86-20 Mar 87 UNAM 1988 22 Jan 88-8 Feb 88 UNAM 1989 24 Jan 89-5 Mar 89 UNAM 1990 25 Nov 89-21 Mar 90 CWR, UABCS, UNAM 1991 28 Nov 90-26 Feb 91 UNAM 1992 26 Dec 91-6 Mar 92 UNAM Baja California Peninsula 1987 10 Feb 87-5 Mar 87 CWR, UABCS 1988 6 Feb 88- 8 Mar 88 CWR, UABCS 1989 23 Jan 89-25 Mar 89 CWR. UABCS 1990 20 Feb 90-31 Mar 90 CWR, UABCS 1991 24 Jan 91-23 Mar 91 UABCS 1992 20 Jan 92-2 Apr 92 UABCS 1993 15 Jan 93-5 Apr 93 UABCS Revillagigedo Archipelago 1986 16 Jan 86-20 Feb 86 UNAM 1987 20 Jan 87-5 Mar 87 UNAM 1988 1 Feb 88-9 Mar 88 UNAM 1989 16 Jan 89-7 Mar 89 UNAM 1990 31 Jan 90-20 Mar 90 UNAM 1991 15 Jan 91-25 Apr 91 UNAM 1992 25 Apr 92-7 May 92 UNAM, UABCS, CS to the population were occurring in the form of ani- mals born in previous years that became identifiable as they gi'ew. Also, some individuals died. Because we would therefore expect to be studying an open population, we chose the Jolly-Seber mark and re- capture model. One of the critical assumptions of this method is the highly unlikely condition that all ani- mals have the same capture probabilities at the mo- ment of the sample (Seber, 1982). If heterogeneity in capture probabilities is present, the open population models will underestimate population size to a greater extent than closed population models (Car- others, 1973; Pollock et al., 1990). Given that there was no way to account for such problems with open population models (Hammond, 1986), we decided to use closed population models to test for variations in capture probabilities. Closed population estimates were obtained and assumptions were tested by using the software pro- gram CAPTURE developed by Otis et al. (1978i, which included an algorithm to select the appropri- ate model after the hypothesis'testing procedure. The conceptual basis for this selection procedure is a tradeoff between precision and bias. If a simple model, such as the null model M^ in Otis et al. ( 1978), is used to estimate parameters from data that vio- lates in any way the assumption of equal capture probability, then significant biases are introduced in parameter estimates and sampling variances will be artificially small. On the other hand, if a more com- plex model is used, such as M,,^ of Otis et al.. ( 1978), that allows capture probabilities to vary with time and among individuals, biases may be reduced but the sampling variance will be greater than it should be. The selection procedure takes into account the individual goodness-of-fit tests performed for specific models on the data and the confrontation of related models (i.e. where one model is a particular case of a general one). The significance levels for all these tests are combined in a standard discriminant analysis and the resulting statistic is standardized so that its value ranges from 0 to 1, 1 being the score that indi- cates the appropriate model (Otis et al., 1978). The basic or null model M^^ can be applied to the general case of multiple recaptures in a closed popu- lation, where all animals have equal capture prob- abilities, and where this probability remains constant in time. The model that allows the capture probabil- ity to vary in time, simultaneously permitting indi- viduals to have unequal capture probability (Chao et al., 1991: model M,^ in Otis et al., 1978) was se- lected for estimation of both stock sizes. In addition. Urban R. et al.: Megaptera novaeangliae in waters off the Pacific coast of Mexico 1021 abundance was also estimated for the coastal stock by using the model that only allows relaxation of the requirement for constant capture probability in time (Darroch, 1958; model M^ in Otis et al., 1978). Model M,[, provides the estimator of population size for a situation where capture probabilities vary in time and among individuals. Model Mj provides the estimate of population size when capture probabilities are the same among individu- als but vary in time (Otis et al., 1978). Outputs from CAPTURE were used only as a "screening technique," as suggested by Menkins and Anderson ( 1988) to investigate departures from the assumption of equal catchability. The test for population closure within CAPTURE was ignoi-ed because its power is low (Otis et al., 1978). Open population estimation was done through the software program RECAP^ This program provides estimates of parameters under the basic Jolly-Seber model where all individuals have equal capture prob- abilities and survivorship, but these were allowed to vary between sampling occasions. Also, it incorpo- rates a modification of the Jolly-Seber model (modi- fied J-S model) that constrains estimates to feasible values, stabilizing them and providing more reliable confidence intervals (Buckland, 1980). Open popu- lation estimates were also obtained by using the soft- ware program JOLLY described by Pollock et al. (1990). Of particular interest is the goodness-of-fit tests performed for this program to investigate how well our data are explained by the Jolly-Seber model or any of the variants included in the program. The results obtained under such variants were also very useful because they allowed for the estimation of more precise parameters when either capture prob- abilities or survivorship ( or both ) were kept constant. Even in the case when the more general Jolly-Seber model was chosen (model A), observation of survi- vorship estimates under model 2 (temporary trap response model ) were useful to look for the amount of transit within the areas occupied by each stock. Results and discussion Photo identification A total of 1184 humpback whales were identified in Mexican waters between 1986 and 1993. Of the to- Table 3 Number of liumpback wha les identified in Mexican water 5 from 1986 to 1993. Region 1986 1987 1988 1989 1990 1991 1992 1993 Coast Animals caught 74 71 74 83 218 109 180 92 Newly caught 74 65 61 67 174 91 140 61 Revillagigedo Animals caught 26 41 73 106 48 189 36 Newly caught 26 37 64 86 33 143 23 Table 4 Estimates of population size lA') with closed population models (see "Mater als an d methods" section for descrip- | tions of the models) Model N 95% CI Capture prob. P Coastal stock M„ 2039 1728-2194 0.05 M,' 1828 1660-2076 0.06 average M.h' 2188 1861-2612 0.05 average Revillagigedo stock M„ 1078 807-1169 0.0687 M, 874 762-1020 0.084 average M.h' 1308 1043-1684 0.057 average ' Selected mod Ms. ■■^ Contact Stephen Buckland. Mathematical Institute. North Haugh, University of St Andrews, St. Andrews, Fife KY169SS, U.K. tal, 733 individuals were observed in the area occu- pied by the coastal stock, 412 in Revillagigedo, and 39 in both areas (a summary of the number of individuals identified in each area is provided in Table 3). Abundance estimation The closed population estimators showed that het- erogeneity and variations in capture probabilities in time are present in both stocks. However, the confi- dence intervals built for models M, and M,,^ overlap to a gi-eater extent in the coastal estimates than in the Revillagigedo estimates, indicating that both models M^^ and M, are equally good for the coastal stock (Table 4). These relatively small capture prob- abilities estimated with the closed models are consis- tent with a positive bias in the population size esti- mates because, as will be discussed, the populations were open through the period of study. Therefore, the closed population models were useful in showing that the assumption of equal catchability is not met. 1022 Fishery Bulletin 97(4), 1999 Table 5 Estimates of population size (A^) with the progi model A ( general ) ( see "Materials and methods descriptions of the models). am JOLLY. ' section for Year N 95% CI Capt ure prob. P Coastal stock' 1987 111 119-1424 0.08 1988 484 215-753 0.14 1989 791 382-1220 0.10 1990 1247 682-1811 0.17 1991 1660 610-2712 0.06 1992 1702 488-2917 0.10 Average 1109 793-1425 0.11 Revillagigedo stock- 1987 134 13-256 0.25 1988 367 125-609 0.18 1989 593 288-897 0.17 1990 456 179-733 0.10 1991 1569 -140-3277 0.12 Average 623 268-980 0.16 ' X^ Goodness-of-fit model A X" Goodness-of-fit model 2 .r = 5.968 P = .v = 1.1.32 P = 0.8754 0.8892 - x' Goodness X^ Goodness of-fit model A of-fit model 2 X = 13.23 P = V = 0.448 P = 0.0667 0.799 The Jolly-Seber analysis indicated that the open population model explained in a better way the data for both Revillagigedo and coastal stocks (Table 5). In the Revillagigedo stock the goodness of fit x' test on model A (general model) did not reject the null hypothesis (P=0.0667), but the result showed that dataarebetter explained (P=0. 799) by model 2 (tem- porary trap response model), assuming that survival rates of newly captured animals differ from those of previously captured animals. In other words, this test indicates that some animals may have moved into the capture site and once captured, they could have left and even re-entered the sampling area. This in- ference an be supported by the fact that data for this stock were collected only in Socorro Island and that some animals may have a preference for anther is- land, such as Clarion, far away from the sampling area (Fig. 1). For the coastal stock, the goodness-of- fit x~ tests indicate that the model A (P=0.87.54 ) and model 2 (P=0.8892) are almost equally good in ex- plaining the data. Although sampling coverage was wider for this stock, allowing a better collection of photographs for moving anirfials, this problem still remained, but had a smaller effect on the estimates. Having accepted that the Jolly-Seber model ex- plained the data well for the two stocks, population- size estimates were obtained with the software pro- Table 6 Estimates of population size (A') with the program RECAP, modified Jolly-Seber model (see "Materials and methods" section for descriptions of the models). Year N 95% CI Capture prob. P Coastal stock 1987 822 425-1142 0.09 1988 580 327-931 0.13 1989 825 519-1229 0.10 1990 1120 824-1613 0.19 1991 1813 1168-2686 0.06 1992 1813 918-2505 0.10 Average 1162 889-1406 Revillagigedo stock 1987 167 49-484 0.25 1988 414 215-691 0.18 1989 610 388-825 0.17 1990 606 359-891 0.08 1991 914 590-1193 0.21 Average 642 373-677 gram RECAP. Results show that the modified esti- mator (modified J-S model) provided more stable estimates and narrower and more realistic confidence intervals (Table 6). Because capture probabilities are larger than those obtained by the closed population methods, and con- sidering that the Jolly-Seber estimates may be nega- tively biased given heterogeneity, population size are likely overestimated by the closed population methods. In light of the above arguments, we conclude that the best estimates of population size for the Mexi- can stocks, using the data at hand, are those pro- vided by the modified Jolly-Seber model of program RECAP. Estimates obtained with closed population methods are not recommended when data are pooled for the whole extension of the study; however, it may be worth obtaining estimates with pairs of years with models testing for heterogeneity and capture prob- abilities varying with time, although certain infor- mation and precision may be sacrificed. Estimates for each year show an apparent trend towards increase for both stocks. Such a trend was considered an artifact of a reduction in the bias caused by heterogeneity in capture probabilities af- ter sample size increased with time. Consequently, we consider that the best estimates for the stock sizes were those obtained for the last years and are 1813 (95^^ CI: 918-2505 ) fbr the coastal stock m 1992, and 914 (95'7r CI: 590-1193) for the Revillagigedo stock in 1991. Urban R et al : Megaptera novaeangliae in waters off tfie Pacific coast of Mexico 1023 Acknowledgments We would like to thank Jeff Laake for valuable ad- vice and fruitful discussions. Judith Zeh, Jay Barlow, Phil Clapham, and two anonymous reviewers pro- vided helpful suggestions for the improvement of the paper. This work was supported by the Direccion General de Investigacion Cientifica y Superacion Academica (DGICSA), C88-01-0414 (1988-90), Earthwatch (1988-90), and the Consejo Nacional de Ciencia y Tecnologia, CONACyT (1991-92). We worked under the permits of the Institute Nacional de Ecologia, SEDUE, SEMARNAP. Literature cited Alvarez, F. C. 1987. Fotoidentificacion del Rorcual jorobado {Megaptera novaeangliae. Borowski, 1781), en las aguas adyacentes a Isla Isabel, Nayarit, Mexico (Cetacea: Balaenopteridae). Professional thesis, Facultad de Ciencias, UNAM, 107 p. Alvarez, C. A., Aguayo L., R. Rueda, and J. Urban R. 1990. A note on the stock size of humpback whales along the Pacific coast of Mexico. Rep. Int. Whal. Comm. Spec. Issue 12:191-193. Baker, C. S., and L. M. Herman. 1987. Alternate population estimates of humpback whales (Megaptera novaeangliae) in Hawaiian waters. Can. J. Zool. 65:2.818-2,821. Baker, C. S., L. M. Herman, A. Perry, W. S. Lawton, J. M. Straley, and J. H. Straley. 1985. Population characteristics and migration of summer and late-season humpback whales (Megaptera novaeangliae ) in southeastern Alaska. Mar. Mamm. Sci. 1:304-323. Baker, C. S., L. M. Herman, A. A. Wolman, H. E. Winn, J. Hall, G. Kaufman, J. Reinke, and J. Ostman. 1986. The migratory movement and population structure of humpback whales (Megaptera novaeangliae} m the cen- tral and eastern North Pacific. Mar. Ecol. Prog. Ser. 31:105-119. Buckland, S. T. 1980. A modified analysis of the Jolly-Seber capture-recap- ture model. Biometrics 36:419-435. Calambokidis, J., J. C. Cubbage, G. H. Steiger, K. C. Balcomb, and P. Bloedel. 1990. Population estimates of humpback whales in the Gulf of the Farallones. California. Rep. Int. Whal. Comm. Spec. Issue 12:343-48. Campos R., R. 1987. Fotoidentificacion y comportamiento del Rorcual jorobado, Megaptera novaeangliae (Borowski. 1781). en las aguas adyacentes al Archipielago de Revillagigedo. Mexico. (Cetacea: Balaenopteridae). Professional thesis. Facultad de Ciencias, UNAM. 134 p. Carlson, C. A., C. A. Mayo, and H. Whitehead. 1990. Changer in the ventral fluke pattern of the hump- back whale, Megaptera novaeangliae, and its affect on matching. Rep. Int. Whal. Comm. Spec. Issue 12:10.5-112. "Carothers, A. D. 1973. The effects of unequal catchability on Jolly-Seber estimates. Biometrics 29:79-100. Cerchio, S. 1998. Estimates of humpback whale abundance off Kauai, Hawaii, 1989 to 1993: evaluating biases associated with sampling the Hawaiian Islands breeding assemblage. Mar Ecol. Prog. Ser. 175:23-34. Cerchio, S., C. M. Gabriele, T. F. Norris, and L. M. Herman. 1998. Movements of humpback whales between Kauai and Hawaii: implications for population structure and abun- dance estimat'on in the Hawaiian Islands. Mar. Ecol. Prog. Ser. 175:13-22. Darling, J. D., K. M. Gibson, and G. K Silber. 1983. Observations on the abundance and behavior of humpback whales (Megaptera novaeangliae) off West Maui, Hawaii, 1977-79. In R. Payne (ed.l. Communication and behavior of whales, p. 201-22. Westview Press, Boulder. Darling, J. D., and H. Morowitz. 1986. Census of "Hawaiian" humpback whales (Megaptera novaeangliae) by individual identification. Can. J. Zool. 64:105-111. Gambel, R. 1976. World whale stocks. Mar Rev. 6:41-53. Hammond, P. S. 1986. Estimating the size of naturally marked whale popu- lation using capture-recapture techniques. Rep. Int. Whal. Comm. Spec. Issue 8:253-282 Jaramillo L., A. 1995. Fotoidentificacion del Rorcual jorobado (Megaptera novaeangliae. Borowski, 1781), en las aguas adyacentes a Isla Isabel, Nayarit, Mexico. (Cetacea; Balaenopteridae). Professional thesis, Facultad de Ciencias, UNAM, 107 p. Johnson, J. H., and A. A. Wolman. 1984. The humpback whale, Megaptera novaeangliae. Mar Fish. Rev 46:30-37. Katona, S. K., and H. P. Whitehead. 1981. Identifying whales using their natural markings. Polar Rec. 20:439-444. Ladron de Guevara, P., J. Urban R., M. Salinas Z., J. Jacobsen, K. C. Balcomb, A. Jaramillo L., D. Claridge, and A. Aguayo L. 1993. Relationships among winter aggregations of hump- back whales. Megaptera novaeangliae. in the Mexican Pacific. In Abstracts of the XVIII Reunion Internacional para el Estudio de los Mamiferos Marines, La Paz, Mexico, May 4-7, 1993. p, 26. Medrano-Gonzalez, L., L. Aguayo L., J. Urban R., and C. S. Baker. 1995a. Diversity and distribution of mitochondrial DNA lin- eages among humpback whales. Megaptera novaeangliae, in the Mexican Pacific. Can. J. Zool. 73:1.73.5-1.743. Medrano-Gonzalez, L. J. Urban R., and C. S. Baker. 1994. Sex and maternal lineage identities of humpback whales in the Mexican Pacific. In Abstracts of the Int. Symposium of Marine Mammal Genetics, La Jolla, CA. 23- 24 September 1994. 1995b. Short and long term population structure of hump- back whales in the eastern North Pacific: the definition of management units and evolutionary significant units revisited. In abstracts of the XX Reunion Internacional para el Estudio de los Mamiferos Marines, La Paz. Mexico. 18-22 April 1995, p. 35. Menkins, G. E., Jr., and S. H. Anderson. 1988. Estimation of small-mammal population size. Ecology 69:1,952-1,959. Otis, D. L., K. P. Burnham, C. G. White, and D. R. Anderson. 1978. Statistical inference from captures data on closed animal populations. Wildl. Monogr. 62, 135 p. 1024 Fishery Bulletin 97(4), 1999 Pollock, K. H., J. D. Nichols, C. Brownie, and J. E. Hines. 1990. Statistical inference for capture-recapture experi- ments. Wildl. Monogr. 107. 97 p. Rice, D. W. 1974. Whales and whales research in the Eastern North Pacific. /;; W. E. Schevill led.), The whale problem, p. 170- 195. Harvard. Univ. Press, Cambridge, MA. 1978. The humpback whale in the North Pacific: distribu- tioQ^ exploitation, and numbers. In K. S. Norris and R. Reeves (eds.), Report on a workshop on problems related to humpback whales iMegaptera novaeangliae) in Hawaii, p. 29-44. U.S. Marine Mammal Commission, Washing- ton. D.C. Seber, G. A. F. 1982. The estimation of animal abundance and related parameters, 2nd ed. MacMillan, New York, NY, 654 p. Urban R., J., and A. Aguayo L. 1987. Spatial and seasonal distribution of the humpback whale, Megaptera novaeangliae. in the Mexican Pacific. Mar Mamm. Sci. 3(4):333-344. Urban R., J., A. Aguayo, L., M. Salinas, Z., R. Campos, R., K. C. Balcomb, J. K. Jacobsen, P. Ladron de G. and C. Alvarez F. 1989. Abundance and interactions of the humpback whale, in the Mexican breeding grounds. In Abstracts of the 8th Biennial Conference on the Biology of Marine Mammals. Monterey. CA. Urban R., J., K. C. Balcomb, C. Alvarez, P. Bloedel, P. Cubbage, J. Calambokidis, J. Steiger, and A. Aguayo L. 1987. Photo-identification matches of humpback whales iMegaptera novaeangliae) between Mexico and central California. In Abstracts of the 17"' Biennial Conference on the Biology of Marine Mammals. Miami, FL. ! 1025 Abstract.— A two-ship sur\-ey of the I'astern tropical Pacific collected sur- face plankton samples during El Nino in August-November 1987. Although the diversity of cephalopods in these samples was low, cephalopod abun- dance was extremely high. Most of the cephalopods collected were rhyncho- teuthion paralarvaeof an ommastrephid squid, either Doscidicus gtgas or Stheno- tciitlus oualaniensis. High abundances (hundreds of paralarvae in a 15-min tow) were concentrated in a band par- allel to the coast, 740-900 km off Cen- tral America. This band of high abun- dance was approximately coincident with the 29 = C surface isotherm. Maxi- mum abundance (>12.000 rhyncho- teuthions in a 15-min tow) was four orders of magnitude greater than back- ground levels and an order of magni- tude greater than any other report of cephalopod abundance. Based on flow- meter readings from the sampler, the anomalous abundance was not a sam- pling artifact. Size-frequency analysis indicates that this patch cannot be ex- plained as a result of recent hatching from an egg mass. This abundance may have resulted from warm El Nino wa- ters, aggregation by convergence of sur- face currents, or the interaction of these factors during the squid's spawning season. Extraordinary abundance of squid paralarvae in the tropical eastern Pacific Ocean during El Nino of 1987 Michael Vecchione Systematics Laboratory National Marine Fisheries Service National Museum of Natural History Washington, DC 20560 E-mail address vecchione michaeliSinmnh si edu * Manuscript accepted 26 October 1998. Fish. Bull. 97:1025-1030 ( 1999). Reports of the effects of El Nino/ Southern Oscillation have been widespread in both scientific litera- ture and popular news media. Shifts in distribution and decreases in abundance of marine animals re- sulting from El Nino have received much attention. Cephalopods are important components of the ma- rine food web, as well as the targets of commercial fisheries. I report here on the serendipitous collection of squid paralarvae during El Nino 1987 at abundances an order of magnitude higher than any previ- ously known. Materials and methods A two-ship survey for a visual cen- sus of marine mammals was con- ducted in the eastern tropical Pa- cific Ocean during August-Novem- ber 1987. During four cruise legs, the NOAA Ship David Starr Jordan surveyed the area between the equator— 19°N and 79-121°W, while the NOAA Ship MacArthin\ also in four legs, covered the region westward of the Jordan survey to 148 -W and southward to 11°S. Sur- face plankton samplers (Manta nets; Brown and Cheng, 1981) were towed by both ships almost every day after sunset to collect fish lar- vae. Surface temperature was mea- sured by bucket thermometer con- currently with each plankton tow. The distance of each tow was esti- mated by a flow meter mounted be- low the net. Extremely high num- bers of cephalopod paralarvae were found in some of these samples when they were sorted for ichthyo- plankton after the conclusion of the cruises. Results The diversity of cephalopods in these samples was quite low; only 15 species were identified and >99% of the 15,052 cephalopods collected were a single species of the squid family Ommastrephidae (Table 1). Ommastrephid paralarvae are dis- tinctive and are referred to as rhyn- choteuthions because the two ten- tacles (as opposed to the eight arms) are fused throughout the early life history. Throughout the entire sur- veyed region, abundance of para- larval ommastrephids, measured as number of rhvnchoteuthions collected per 15-min tow, generally was 0-10 squids per tow. Maximum abundance (22 stations with 20->100 rhyncho- teuthions/tow) was found about 740-900 km offshore in a band par- allel to the coast of Central America (Fig. lA). Within this band at three stations during two separate legs, 379. 461, and 12,354 rhynchoteu- thions were collected, respectively. Although most of the stations in this band were sampled by the Jordan, the northeasternmost MacArthur stations were in the eastern end of 1026 Fishery Bulletin 97(4), 1999 Number of specimens per tow 100° Surface temperature Figure 1 NOAA Ship D.S. Jordan cruise 8710. August-November 1987: (Ai sampling locations and iso- lines of abundance for rhynchoteuthion squid paralarvae; (B) surface isotherms (°C). the band. A secondary patch of abundance (five sta- tions with 20-90 rhynchoteuthions/tow) was sampled by the MacArthur farther to the west (9-13°N, 121- 135°W) and appeared to be a westward extension of the band of peak abundance. The primary band of maximum abundance was approximately coincident with the 29 'C surface iso- therm (Fig. IB). Peak abundance was found in the vicinity of a very productive oceanographic upwelling feature known as the Costa Rica Dome (Li et al., 1983). The secondary patch of abundance was in an area with surface temperatures in the mid-28"C range. It may be associated with a westward exten- sion of the Costa Rica Dome, termed the counter- current ridge (Fiedler et al., 1991). A plot of abun- dance versus surface temperature shows a clear peak between 27.5" and 31.0 C (Fig. 2). The three stations with highest abundances were all in areas of 29.0- 29.5°C surface temperatures (see vertical arrows in Fig. 2). Abundance tapered off late in the survey al- though surface temperatures remained above 27.5'C (Fig. 3), within the optimum range indicated by peak abundance. In addition to a biological-oceanographic pattern of abundance, two other alternatives could explain the pattern described here: 1 ) it could represent con- centrated patches of recent hatchlings from isolated egg masses; or 2) it could result from sampling error associated with changes in towing characteristics of the samplers (e.g. speed, duration). To address alternative 1, 1 examined size-frequency distributions based on dorsal mantle lengths (DML), both within individual samples and in the total pooled population. For samples with > 100 specimens, I measured a subsample of 100. 1 separated the sta- tions into dusk tows (within one hour after local sun- Vecchione: Abundance of squid paralarvae during El Niiio of 1987 1027 Table 1 Cephalopods collected by Manta-net surface zoop ankton samples during the two -shipsurvey of the eastern tropical | Pacific, August-Novem ber 1987. Ship Total Jordan MacArth ur Ommastrephidae 14,445 495 14,950 Argonauta sp. 16 36 52 unid. squid 9 3 12 Liocranchia sp. 1 9 10 Abraliopsis sp. 9 0 9 Onykia? sp. 4 2 6 Chiroteuthis sp. 3 0 3 Pterygioteuthis sp. 2 0 2 Chtenopteryx sp. 1 1 2 Enoploteuthis? sp. 1 1 2 Abra/ia? sp. 1 0 Thysanoteuthis rhombus 1 0 Cranchia scabra 1 0 Leachia sp. 1 0 Octopodidae 0 1 Total 14,495 548 15,043 set) and night tows to determine the contribution of diel variabiHty, from either visual avoidance of the net or vertical migration by larger animals, to the size distribution. The overall size distribution was skewed toward the size of newly hatched animals, with the model length 1.0-1.5 mm DML. The size range in dusk samples was 0.5-3.5 mm DML (Fig. 4A). A small second group of larger squid 3.5-8.5 mm DML (modal length ca. 6.0-6.5 mm DML) not present in dusk tows was collected at night in addition to the smaller paralarvae (Fig. 4B). Some stations, such as Jordan station IV-05 (Fig. 4D), included only very small squid 1-2 mm DML, as might be expected from a hatching event. The greatest abundance, however, at station III-62, was characterized by a broader range of lengths (Fig. 4C) with a larger modal size of 2.0-2.5 mm DML. On the basis of analysis of growth rings in statoliths from Hawaiian S. oualaniensis, a 2.5-mm-ML paralarva is ca. 20 days old (Bigelow, 1991 ). I therefore conclude that this very high abun- dance did not result from the net happening upon an egg mass, either just before or just after hatch- ing, but instead an aggregating mechanism must be responsible for this patch. I examined the second alternative by calculating the variability in flow meter revolutions among tows. The mean and standard deviation for this estimate of sampling efficiency were 2771 and 645, respec- tively. None of the three outliers from this distribu- 24 25 26 27 28 29 30 31 32 Surface temperature Figure 2 Rhynchoteuthion squid paralarval abun- dance plotted against surface temperature for NOAA Ship D.S. Jordan cruise 8710. Vertical arrows represent samples with maximum numbers of paralarvae. 32 S 30 |- 29 a 28 S 27 ^ 26 ^ 25 24 220 240 260 280 300 320 340 140 I 120 S100 S" 80 I 60 S 40 Q. ^ 20 0 220 240 260 280 300 320 340 .lulian date Figure 3 Relationships among water temperature, squid abundance, and date for NOAA Ship D.S. Jordan cruise 8710: (A) surface tem- perature plotted against Julian date; (B) rhynchoteuthion squid paralarval abun- dance plotted against Julian date. Verti- cal arrows represent samples with maxi- mum numbers of paralarvae. Hi tion (one high and two very low) collected >20 rhynchoteuthions. All of the samples with abun- dances >40 rhynchoteuthions came from tows with 2612-2947 revolutions. Although this parameter can be used to estimate the volume of water filtered by the sample, I did not do this because assumptions about the orientation of the net mouth with respect to the sea surface are required for surface samples and this orientation can change greatly with sea 1028 Fishery Bulletin 97(4), 1999 Dusk tows Jordan station ni-62 1,5 2 6 3,5 4.5 5 5 6.5 7 5 8 5 Night tows B 0 5 t 5 2.5 3 5 4 5 5 5 6 6 7,5 8.5 Jordan station rV-05 D 0 5 15 2.5 3.5 4.5 5 5 6.6 T b 8 5 Mantle length (mm) Figure 4 Size-frequency distributions for ommastrephid paralai-vae. All specimens were measured from samples that collected <100 rhynchoteuthions. A subsample of 100 specimens was measured from samples containing >100: (A) pooled distribution for all dusk tows (samples collected <1 h after local sunset); (Bl pooled distribution for all night tows (samples collected >1 h after local sunset); (C) lengths of 100 specimens from the station that collected maximum abundance; (D) example of size-frequency distri- bution expected from a station that sampled a recent hatching event. state. The low variability of flow meter counts in(di- cates that observed abundances did not result from sampling artifacts. Discussion Aside from the band of high rhynchoteuthion abun- dance parallel to the coast and coincident with the 29°C isotherm, these collections are not unusual for surface samples of cephalopod paralarvae. The abun- dance of other cephalopods with paralarvae typically found at the surface, such as Argonauta, was not exceptional. Unusually high abundances in these samples were found only in the paralarvae of ommastrephid squids. The Humboldt squid, Doscidicus gigas, is an ommastrephid sufficiently abundant in this area to support commercial fisheries, but descriptions of the paralar\'ae of this species are not adequate to iden- tify them with confidence. However, the general mor- phology, morphometries, and chromatophore patterns of the paralarvae reported here are consistent with those of another widely distributed and abundant ommastrephid, Stheiwteiithis oualaniensis, whose paralarvae have been described from off Hawaii (Harman and Young, 1985). It is not possible at this time to be certain which of these species composed the high abundances reported here. Although natu- ral spawning has not been observed directly for ei- ther of these species, ommastrephids are known to spawn large egg masses that are gelatinous and pe- lagic and are very difficult to collect with nets. Spawning aggregations of S. oualaniensis and other ommastrephids have been located in other areas. It seems reasonable to assume that either species may aggregate in the study area to produce egg masses. No information is available on the abundance of these squid paralarvae in subsurface waters of this region. The highest abundance of S. oualaniensis in Hawaiian waters is found in the mixed layer from the surface to 20 m depth (Young and Hirota, 1998). Surface abundance therefore appears to be a reason- able indication of overall paralarval distribution of this species. Furthermore, the depth of maximum zooplankton abundance in the eastern tropical I Vecchione: Abundance of squid paralarvae during El Nino of 1987 1029 Pacific is generally limited by an oxygen minimum layer beneath the thermocline (Saltzman and Wishner, 1997). Thus, the distribution of these shal- low-living paralarvae may have been compressed toward the surface by the oxygen-minimum layer. Expendable bathythermograph measurements by the Jordan during this period showed the mixed- layer depth to be 20-40 m, except where the ther- mocline was depressed within an anticyclonic eddy located very close to the stations with maximum squid abundance (Hansen and Maul, 1991). Conver- gence of surface currents associated with the downwelling that depresses the thermocline could aggregate surface plankton such as these paralarvae ( Bakun and Csirke, 1998 ). Although the area of maxi- mum abundance was near the Costa Rica Dome, an upwelling feature where the thermocline typically shoals from 60 m to 30 m depth ( Balance at al., 1997 ), thermal topography in this area was anomalously flat during El Nino of 1987 (Fiedler et al, 1992). A few other records of cephalopod distribution and abundance are known from this region. In an exten- sive multicruise survey of the eastern tropical Pa- cific in 1967-68 (Okutani, 1974), few.?, oualaniensis or other ommastrephids were collected. Similarly, in a cruise in this region from October 1969 through February 1970, during which surface zooplankton were sampled, no large numbers of paralarvae were encountered, although the most abundant cephalo- pod family sampled was the Ommastrephidae (Ueynagi and Nonaka, 1993). Typical numbers of cephalopod paralai"vae taken in zooplankton samples worldwide number 0 to per- haps 20 squids per sample, usually about 1-5 squids (Vecchione, 1987). The highest abundance previously reported anywhere of which 1 am aware was almost 1000 loliginid paralarvae in a 15-min surface plank- ton tow in the western North Atlantic (Vecchione et al., 1986). The maximum abundance found in the present study was over an order of magnitude greater than that. The samples reported in this study were collected during the peak of a moderate El Nifio event (McPhaden and Hayes, 1990). Surface temperatures in the area averaged 3.5°C warmer than during the same period of the following year (Fiedler et al., 1992). The temperature distribution of these samples indicates that high abundances of rhynchoteuthions would not be expected at temperatures <27.5°C. Therefore, warm El Nino waters probably were the primary reason for the high abundance of squid paralarvae found here. These abundant surface- dwelling paralarvae may have been concentrated to extraordinary densities by convergent surface cur- rents in the vicinity of an anticyclonic eddy. Seasonal occurrence is likely important, because surface tem- peratures remained high until the end of these cruises, whereas squid abundance decreased dra- matically during the final 30 days. The sampling for- tunately was conducted during the season when these squid were hatching. Acknowledgments These samples and data were collected by biologists from the Marine Mammal Program of the National Marine Fisheries Service Southwest Fisheries Sci- ence Center (SWFSC), primarily by R. Pitman and J. Carretta. The samples were sorted in R. Charter's laboratory at SWFSC, and he and G. Moser, also of SWFSC, sent the cephalopods to me. I was assisted in cephalopod sorting and measuring by K. Jackson. The following people provided helpful comments on a draft of this paper: R. Young, G. Moser, B. Collette, and C. Roper. In addition, R. Young graciously pro- vided unpublished data on vertical distribution of Ha- waiian paralai-vae for an early draft of this paper. Literature cited Bakun, A., and J. Csirke. 1998. Environmental processes and recruitment varia- bility. FAO Fish. Tech. Pap. .376:10.5-124. Balance, L. T., R. L. Pitman, and S. B. Reilly. 1997. Seabird community structure along a production gra- dient: importance of competition and energetic constraint. Ecology 78:1502-1518(1997). Bigelow, K. A. 1991. Age and growth of three species of squid paralarvae from Hawaiian waters, as determined by statolith micro- structures. M.S. thesis, Univ. Hawaii. Honolulu, HI, 78 p. Brown, D. M., and L. Cheng. 1981. New net for sampling the ocean surface. Mar Eeol. Prog. Ser. 5:225-227. Fiedler, P. C, F. P. Chavez, D. W. Behringer, and S. B. Reilly. 1992. Physical and biological effects of Los Nifios in the eastern tropical Pacific. 1986-1989. Deep-Sea Res. .39:199-219. Fiedler, P. C, V. Philbrick, and F. P. Chavez. 1991. Oceanic upwelling and productivity in the eastern tropical Pacific. Limnol. Oceanogr. 36:1834-1850. Hansen, D. V., and G. A. Maul. 1991. Anticyclonic current rings in the eastern tropical Pacific Ocean. J. Geophys, Res. 96:6965-6979. Harman, R. F., and R. E. Young. - 1985. The larvae of ommastrephid squids (Cephalopoda, Teuthoidea) from Hawaiian waters. Vie Milieu 35:211-222. Li, W. K. W., D. V. Subba Rao, W. G. Harrison, J. C. Smith, J. J. Cullen, B. Irwin, and T. Piatt. 1983. Autotrophic picoplankton in the tropical ocean. Science (Wash., D.C.) 219:292-295. McPhaden, M. J., and S. P. Hayes. 1990. Variability in the eastern equatorial Pacific Ocean during 1986-1988. J. Geophys. Res. 95:13195-13208. 1030 Fishery Bulletin 97(4), 1999 Okutani, T. 1974. Epipelagic decapod cephalopods collected by micronekton tows during the EASTROPAC expeditions, 1967-1968 (systematic part). Bull. Tokai Reg. Fish. Res. Lab. 80:29-118. Saltzman, J., and K. F. Wishner. 1997. Zooplankton ecology in the eastern tropical Pacific oxygen minimum zone above a seamount: 1. General trends. Deep-Sea Res. 44:907-930. Ueynagi, S., and H. Nonaka. 1993. Distribution of ommastrephid paralarvae in the cen- tral-eastern Pacific Ocean. In T. Okutani, R.K. O'Dor, and T. Kubodera (eds. ), Recent advances in cephalopod fisher- ies biology, p. 587-589. Tokai Univ. Press, Tokyo. Vecchione, M. 1987. Juvenile ecology. In P. Boyle (ed.). Cephalopod life cycles, vol, 2. p. 61-84, Academic Press, London. Vecchione, M., C. F. E. Roper, C. C. Lu, and M. J. Sweeney. 1986. Distribution and relative abundance of planktonic cephalopods in the western North Atlantic. Am. Malac. Bull. 4:101. Young, R. E., and J. Hirota. 1998. Review of the ecology of Sthenoteuthis oualaniensis near the Hawaiian Archipelago. /;; T Okutani ( ed. ), Large pelagic squids, p. 131-143. Japan Marine Fishery Re- sources Research Center, Tokyo. 1031 Abstract. -The whitetip nyingfish, Cheilopogon xenopterus. is an epipe- lagic resident of tropical and subtropi- cal eastern Pacific waters. Its eggs are spherical, average 1.8 mm in diameter, and have an homogeneous yolk and no oil globule. About 53 filaments averag- ing 1 mm in length are evenly distrib- uted on the chorion. The notochord flexes, fin-ray formation is nearly com- plete, and the characteristic larval pig- mentation pattern is established prior to hatching at a larval length of about 2.8-3.3 mm. Larvae hatch with pig- mented eyes, functional mouth, and little remaining yolk. Pectoral- and pel- vic-fin rays initially are short but elon- gate rapidly to ca. 25-50^7^ and 20-40'7f of body length, respectively. A pair of mandibular barbels form at about 4 mm and fuse mesially at about 8 mm. Scales begin to form along the lateral line at about 13-14 mm and cover the body by 26 mm. The characteristic pigment pattern, visible through the early juvenile stage, consists of the following: melanophores scattered over the mid- and hindbrain. continuing posteriorly as two rows (in- creasing to four or more rows ) along the dorsal margin; a row of melanophores on the horizontal septum of the tail (af- ter hatching); a patch on each side over the hypural area; and two rows along the anal-fin base. Internal pigment is present on the mid- and hindbrain. over the gut. and over the notochord. The pectoral and pelvic fins are sparsely pigmented at hatching and become in- creasingly pigmented with growth. A barred pigment pattern begins to develop on the body at about 8 mm and by the juvenile stage about six bars are present. Early life history stages of the whitetip flyingfish, Cheilopogon xenopterus (Gilbert, 1890) (Pisces: Exocoetidae) William Watson Southwest Fisheries Science Center National Marine Fisheries Service. NOAA PO Box 271 La Jolla, California 92038-0271 E-mail address billwatsonia'ucsd edu Manuscript accepted 20 November 1998. Fish. Bull. 97:1031-1042 ( 1999). Flyingfishes, surface-oriented resi- dents of all warm oceans, are well known for their ability to leap from the water and glide over long dis- tances. Flyingfishes can be abundant locally, many are attracted to light, and the gustatory quality of their flesh is generally good (Heemstra and Parin, 1986; Gillett and lanelli, 1991; Parin, 1995). Directed fisheries for flyingfish currently exist, primarily in parts of the Indo-West Pacific and Caribbean; few such fisheries exist elsewhere ( Gillett and lanelli, 1991; Oxenford et al., 1995; Parin, 1995). The family contains eight or nine genera and about 50-60 species (Nelson, 1994; Dasilao et al., 1996); nearly half the species are in the genus Cheilopogon (Heemstra and Parin, 1986). Eggs have been de- scribed for nine Cheilopogon species (Barnhart, 1932; Hubbs and Kampa, 1946; Miller, 1952; Imai, 1959; Gorbunova and Parin, 1963; Parin and Gorbunova, 1964; Kovalevskaya, 1965; Gibbs and Staiger, 1970; Vijayaraghavan, 1975; Shigonova and Kovalevskaya, 1991; Watson, 1996), and at least some larval stages are known for 16 species (Hildebrand and Cable, 1930; Barnhart, 1932; Breder, 1938; Hubbs and Kampa, 1946; Imai, 1959, 1960; Gorbunova and Parin, 1963; Kovalevskaya, 1965, 1975, 1977, 1982; Vijayarhagavan, 1975; Chen, 1987, 1988; Shigonova and Kovalevskaya, 1991; Belyanina, 1993; Parin and Belyanina, 1996; Watson, 1996). The purposes of this paper are to provide a description of the egg, larval, and early juvenile stages of the endemic eastern Pacific species Cheilopogon xenopterus (Gil- bert, 1890) and to compare these briefly with the early stages described for other Cheilopogon species in the eastern Pacific. Materials and methods Descriptions are based on 20 eggs and a size series of 45 larvae (2.8- 22.8 mm) and 8 juveniles (25.9- 44.8 mm). Neustonic eggs and lar- vae and three of the juveniles were collected with Manta nets (Brown and Cheng, 1981) during Marine Mammals Division (Southwest Fisheries Science Center) dolphin surveys in the eastern tropical Pa- cific (Thayer et al., 1988a, 1988b; Lierheimer et al., 1989, 1990; Phil- brick et al.. 1991, 1993 ). The Manta samples were taken nightly from late July through early November or December, 1987-90, and 1992, between about 2-16°N and east of about 115°W. A few juvenile and adult specimens of C atrisignis, C. dorsomaciilata, C. furcatus, C. papilio, C. spilonotopterus, and C xenopterus, radiographed to make fin-ray and vertebral counts for comparison with counts from series specimens, were obtained from the Scripps Institution of Oceanogra- phy Marine Vertebrates Collection (SIO). 1032 Fishery Bulletin 97(4), 1999 Eggs, larvae, and juveniles (except those used ex- clusively for radiography) were measured to the near- est 0.04 mm by using a Wild M-5 binocular micro- scope equipped with an ocular micrometer. Dimen- sions measured for eggs included chorion diameter, maximum yolk diameter, and lengths of the shortest and longest intact chorionic filaments, and for lar- vae andjuveniles, body length (BL), preanal length (PAL), head length (HL), snout length (SnL), barbel length (BbL), eye diameter (ED), head width (HW), body depth (ED), and lengths of the pectoral (PjL) and pelvic (P2L) fins. All these dimensions, except chorionic filament length (measured along the axis of the filament from its point of attachment on the chorion to its free tip), are defined by Moser (1996). Larval lengths refer to formalin-preserved body length. All descriptions of pigmentation refer solely to melanistic pigment. One late embryo (3.3 mm) dissected from the egg and eight larvae (2.9-16.8 mm) were cleared and stained with alcian blue and alizarin red S accord- ing to the method of Taylor and VanDyke (1985) to elucidate the development of the axial skeleton and fins and to aid in making counts. Illustrations were made with a Wild M-5 microscope equipped with a camera lucida. Identification Specimens were identified by the series method. A size series of larvae andjuveniles, linked by shared morphological, meristic, and pigmentation charac- ters, was traced down to recently hatched larvae from juveniles that could be identified by using known juvenile and adult characters. The smallest larvae shared a pigmentation pattern with late-stage em- bryos that allowed identification of the eggs. Of the eight fiyingfish genera in the eastern tropi- cal Pacific (Parin, 1995; Dasilao et al., 1996), only Cheilopogon has paired mandibular barbels during at least part of the larval and juvenile stages (Parin, 1961a, 1961b; Collette et al., 1984). {Parexocoetus brachypterus develops a pair of mandibular barbels, and a small beak as well, during the juvenile stage: Collette et al., 1984.) Of the nine Cheilopogon spe- cies in and near the area (Table 1; Parin, 1995), lar- vae are known for five: (C. atrisignis: Chen, 1987, 1988; C. furcatus: Hildebrand and Cable, 1930; C. heterurus hiibbsi: Barnhart, 1932; Watson, 1996; C. pinnatibarbatus californicus: Hubbs and Kampa, 1946; Watson, 1996; and C. spilonotopterus: Kova- levskaya, 1977; Chen, 1987, 1988). All five differ from the series treated here. Cheilopogon rapanouiensis was eliminated from consideration by its higher ver- tebral, pectoral-fin ray, and predorsal scale counts (45-46 vs. 43, 16-17 vs. 13-15, and 31-33 vs. 27-28, respectively). Cheilopogon papilio was eliminated because of its lower dorsal-fin ray count (9-10 vs. 12-13), usually lower pectoral-fin ray count (12-13 vs. 13-15), because of separate rather than fused barbels and much heavier pectoral- and pelvic-fin pigmentation in the juvenile stage, and because of restricted distribution in coastal waters and the lower Gulf of California (Parin, 1995). The remaining two species, C. dorsomaculata and C. xenopterus, are widely distributed in the eastern tropical Pacific (Parin, 1995), are quite similar in appearance as small (9 mm; 13-15 rays) match the range for C. xenopterus and fall below to within the lower half of the range for C. dorso- maculata (Table 1). Juvenile C. dorsomaculata typi- cally have less pelvic-fin pigment than C. xenopterus, and in this character the larger series specimens are consistent with C. xenopterus. Thus, 1 concluded that the series is C. xenopterus. Specimens examined Listings are given as cruise and station number or SIO catalogue number, and in parentheses number of specimens and size range. Specimens listed with cruise numbers are housed at the National Marine Fisheries Service Southwest Fisheries Science Cen- ter; specimens with SIO numbers are housed at the Scripps Institution of Oceanography Marine Verte- brates Collection. Cheilopogon xenopterus (Gilbert, 1890). Eggs: 8910JD: 3-100 (3: 1.8-1.9 mm), 3-102 (7: 1.8- 1.9 mm), 4-107 (1: 1.7 mm); 9210JD: 1-001 (4: 1.8- 1.9 mm), 1-008 (4: 1.8-1.9 mm), 3-047 (1: 1.8 mm). Larvae; 8710JD: 1-009 ( 1: 15.2 mm), 2-016A ( 1: 2.8 mm), 3-046 (6: 3.4-4.8 mm), 3-054 (1: 2.9 mm), 3- 056 (1: 4.9 mm), 3-075 ( 1: 2.9 mm), 3-078 (2: 5.3, 14.3 mm); 8710M4: 1-003 (1: 4.1 mm); 8810JD; 2-049(1: 8.0 mm), 3-086 (1: 7.5 mml, 3-118 (5: 8.7-13.6 mm), 4-153 ( 1: 8.3 mm); 8910JD: 1-015 ( 1: 2.9 mm), 1-019 (1:6.4 mm), 1-021(1: 13.7 mm), 2-056 (1: 6.8 mm), 3- 104(1: 4.8 mm); 8910M4: 1-010(1: 12.5 mm), 1-012 (1: 3.4 mm), 4-143 (2: 9.8, 13.9 mm); 9010JD: Manta 54(1: 13.4 mm), Manta 65(1: 6.4 mm), Manta 74(1: 22.8 mm); 9210JD: Manta 2 (1: 8.1 mm), Manta 9 Watson: Early life history stages of Cheilopogon xenopterus 1033 Table 1 Selected meristic characters for the Cheilopog on species that occur in and near the eastern tropical Pacific Ocean. The pectoral- | fin ray count includes the small, spine-like first pectoral -fin ray. Data are from Parin ( 1960 1961a), Watson (1996), and counts made during this study. D = dorsal; A = anal; Pi = pectoral; Proc.C = procurrent caudal-fin rays. Vertebrae Fin rays Predorsal Species Abdominal Caudal Total D A P. Proc. C scales atrisignis 28-30 15-16 43-45 13-15 9-10 13-15 6-1-8 31-40 dorsomaculata 28-29 14-15 42-44 11-14 9-10 14-17 5-6-f6-7 23-25 furcatus 29-31 14-15 43-46 12-14 9-11 14-18 7-H6-8 27-35 heterurus hubbsi 31-33 15-17 47-49 12-14 8-11 14-16 6-7-1-7 28-36 papilio 28-29 14-15 42-44 9-10 9-10 12-13 4-5-h6 29-33 pmnatibarbatus californicus 31-35 16-17 48-51 9-13 9-12 14-15 5-6-1-6-8 40-43 rapanouiensis 30-31 15-16 45-46 11-12 10-11 16-17 31-33 spilonotopterus 29 14 43 13-14 10-11 13-15 6-1-7 29-34 xenopterus 27-30 13-15 42-44 10-13 9-10 13-15 5-6-1-6-8 27-30 (1: 6.3 mm), Manta 11 (2: 4.7, 5.3 mm), Manta 19(?) (1: 8.1 mm), Manta 34 (2: 5.3, 5.6 mm), Manta 37 ( 1: 16.5 mm), Manta 47 (1: 6.5 mm), Manta 51 ( 1: 6.8 mm); SIO 63-54(1: 18 mm). Juveniles: 8710JD: 1-006 (1: 25.9 mm); 8810JD: 3- 108 (1: 44.8 mm); 8910JD: 1-039 (1: 36.4 mm); SIO 63-31 (1: 116 mm); SIO 63-96 (1: 38.5 mm); SIO 63- 105 (7: 34.5-57.0 mm); SIO 63-608 (3: 64-92 mm); SIO 64-539 (1: 62 mm); SIO 72-121 (2: 42, 45 mm); SIO 73-400 (3: 63-76 mm). Cheilopogon atrisignis (Jenkins, 1903). SIO 79-29 (1: 52 mm). Cheilopogon dorsomaculata (Fowler, 1944). SIO 52- 399 (1: 46 mm); SIO 52-416 (6: 147-214 mm); SIO 58-318 (1: 179 mm); SIO 78-214 (1: 36.5 mm). Cheilopogon furcatus (Mitchill, 1815). SIO 76-246 (1: 52 mm); SIO 93-89 (1: 183.5 mm FL). Cheilopogon papilio (Clarke, 1936). SIO 58-395 (2: 65-75 mm); SIO 69-387 (2: 20, 21 mm); SIO 93-95 (2: 109, 115.5 mm). Cheilopogon spilonotopterus (Bleeker, 1866). SIO 60-265 (1: 103 mm); SIO 69-405 (1; 39 mm). Description of eggs Morphology Eggs spherical, average 1.8 mm in diameter (range 1.7-1.9 mm), have narrow perivitelline space (mean yolk diameter 1.7 mm, range 1.5-1.8 mm), homoge- ^neous yolk, and lack oil globules (Fig. 1). Between about 42 and 64 slender filaments (mean 53) evenly distributed over smooth, transparent chorion; each attached to chorion at one end and all of similar length (mean 1.2 mm, range 0.7-1.4 mm). Pigmentation Chorion unpigmented, yolk colorless to pale yellow. Embryonic pigmentation first appears about midway through development (stage VIII; embryonic tail length increases fi-om 507? to 100'^ HL during this stage; e.g. Moser and Ahlstrom, 1985) as one to few small melanophores dorsolaterally on midbrain area and small cluster just anterior to each pectoral-fin bud (Fig. IB). Pigmentation increases dorsolaterally on head, forms around margins of eyes and ventrally on head in stage IX (tail > HL but < 507c yolksac length). During this stage melanophores spread pos- teriorly and dorsally from prepectoral clusters, form- ing two rows on dorsum to midway along embryonic axis, and melanophores form dorsally on developing gut, beginning posteriorly early in stage. By stage X (tail extends 50-75^^ of yolksac length) characteris- tic larval pigment pattern clearly visible, with dor- sal rows extending to near end of tail, a few melano- phores around margin of caudal peduncle, and two rows along base of anal fin (Fig. IC). Near end of embryonic development (stage XI: tail extends >75% of yolksac length) melanophores fill in over central axis of head, form near distal margins of pectoral fins and proximally on caudal fin, may form near distal margins of pelvic fins (present in one of three stage-XI specimens examined), form in internal se- ries over notochord and posteriorly under notochord, and become increasingly dense on dorsal and ven- tral margins near midtail. Eyes become fully pig- mented during stage XI. 1034 Fishery Bulletin 97(4), 1999 Figure t Eggs of Cheilopogon xenoptenis. (A) stage VI, 1.8 mm (8910JD. station 3-1021; (B) stage VIII, 1.8 mm (9210JD, station 1-008); (C) stage X, 1.9 mm (8910JD, station 3-100). Description of larvae Morphology Larvae hatch at about 2.9-3.3 mm length and have Httle remaining yolk, functional mouth, fully flexed notochord, and rays forming in all fms (Fig. 2). Lar- vae moderately elongate, with preanal length near 807c BL at hatching, become increasingly elongate with preanal length decreasing to near lO^c BL by ca. 8 mm (Table 2). Eyes initially oval (vertical axis about 707( horizontal axis), gradually becoming nearly round. Pectoral- and pelvic-fm rays initially short (near 10% BL), rapidly elongate to ca. 25-50^^ and 20-409f BL, respectively. Height of dorsal fin increases from about 15% BL in large lai'vae ( IS- IS mm) to about 28% BL in small juveniles (by ca. 36 mm). A pair of mandibular barbels forms at about 4 mm. Barbels originate as low, anteroventral thick- ening that elongates into a slender, flattened flap on each side of lower jaw. Barbels broaden, develop frilled margins, and fuse mesially at their bases by about 8.3 mm (Figs. 2-4). Scales form along lateral line beginning at about 13-14 mm and cover body by 26 mm. There are 27-30 predorsal scales and seven scale rows between dorsal-fin origin and lateral line. Vertebral column and fin development Notochord flexion begins during embryonic stage VIII (dorsal, anal, caudal finfolds first visible in this stage) and is completed early in stage IX. Vertebral column ossification begins soon after hatching. Neural and haemal arches and spines apparently form first, be- fore corresponding vertebral centra. Ossification of arches apparently anterior to posterior: in 3.8-mm specimen all arches ossifying except last three neu- ral arches, last haemal arch, and last three or four neural and haemal spines (all present as cartilage). Direction of ossification of individual arches appar- ently distad from base: in 3.8-mm specimen haemal arches 11-13 appear to be ossifying from near base Watson: Early life history stages of Cheilopogon xenopterus 1035 Table 2 Summary of measurements oi Cheilopogon xenopterus, expressed as percentage of body length (BL) or head length (HL). For each measurement the mean is given above and the range is given below. For eye diameter (ED), eye length is given first and eye height second; n = number of specimens. Specimens 24.1 mm and larger were considered juveniles. PAL= preanal length; PjL = pectoral-fin length; PjL = pelvic-fin length HW = head width; SnL = snout length; BbL = barbel length. BL(mml n PAL/BL BD/BL HL/BL P,L/BL P.L/BL HW/HL SnL/HL ED/HL BbL/HL 2.8-4.0 9 76 24 33 28(n=6) 28(;i=5) 69 14 42.31 0.3 71-88 22-28 30-39 12-35 18-32 67-74 12-17 39-47, 28-34 0-3 4.1-6.0 10 72 21 30 38(n=8) 38(/i=9» 69 17 41,33 4(n=9) 68-75 19-24 29-33 27-50 34-44 66-75 13-23 38-45, 29-35 2-10 6.1-8.0 8 70 18 26 37 (n=5l 30 67 17 41,34 17 (n=7l 68-73 17-20 24-28 35-40 28-33 63-73 14-20 38-42,31-37 10-24 8.1-10.0 6 69 17 24 39 (/I =5) 35(;!=5) 66 17 43,36 36 66-71 16-18 2.3-25 35-43 30-39 61-72 16-19 40-46, 33-42 24-48 10.1-12.0 2 68 17 22 40(n = ll 31 67 16 44,37 46 68-68 16-17 22-22 30-32 67-67 16-17 43-44 39-53 12.1-14.0 .5 68 15 20 42 35 71 18 46,42 49 66-68 14-16 19-20 39-46 34-37 68-73 16-19 43-48, 39-45 39-57 14.1-16.0 2 73 16 21 48 39 73 15 47,43 65 72-75 16-17 20-21 46-49 37-41 70-76 1.3-17 44-50, 42-43 59-72 16.1-18.0 2 68 16 20 50 41 74 20 45,43 71 68-68 15-18 19-20 49-52 41-41 73-75 17-22 44-47. 41-45 59-83 22.1-24.0 1 68 15 20 58 39 71 20 46,45 61 24.1-26.0 1 69 15 19 56 39 76 18 48,47 55 34.1-36.0 1 71 16 21 65 43 70 24 38,37 43 .36.1-38.0 1 72 16 20 68 44 73 17 42,44 44 38.1-40.0 1 74 15 22 72 43 67 18 36,36 43 40.1-42.0 2 72 16 22 67 42 67 21 39,40 46 71-73 15-16 22-22 67-68 41-42 66-68 20-23 36-41, 38-42 39-53 42.1-44.0 1 71 16 22 69 43 66 20 38,37 39 44.1-46.0 1 77 19 21 72 42 79 19 43,44 53 of each haemapophysis. In 6.3-nim specimen last four neural arches and spines broadening and by 8.2 mm last six much broader than more anterior neural arches and spines. These broad posterior neural arches and spines provide the necessary attachment surfaces for the large supracarinalis posterior, flexor dorsalis, and flexor ventralis muscles (Winterbottom, 1974; Dasilao et al., 1996) which are involved in gener- ating the strong caudal thrust required for gliding. Ossification of each vertebra begins ventrally. As bone spreads dorsad from ventrum, ossifications also form at bases of the two neurapophyses and spread mesially and ventrad to complete centrum. Centra apparently added from anterior to posterior, but uro- style and penultimate centrum ossify before other — caudal vertebrae. All vertebrae ossified by 5.2 mm (Table 3). There are 42-44 vertebrae: 27-30 abdomi- nal and 13-15 caudal (modally 29 -i- 14 = 43). Pleural ribs first visible in 6.3-mm specimen (Table 3 ). These ossify adjacent to parapophyses beginning at third vertebra (at second vertebra on one side in one specimen) and are added posteriorly. Epineural intermuscular bones begin ossifying in myosepta anteriorly, initially remote from vertebrae, by 8.2 mm (Table 3), and pairs added posteriorly. Ossification of each rib and epineural intermuscular bone is both mesial toward adjacent parapophysis, and distal. Full complements of both series not yet formed in largest cleared and stained specimen. Caudal and pectoral are first fins to begin form- ing. Hypural elements first visible in late stage-VIII embryos, begin ossifying soon after hatching (by 2.9 mm). Retrorse basal spur forms on hypural 1 by 6.3 mm. Uroneural ossifying by 5.2 mm and epurals 1 and 2 by 6.3 mm. Epural 3 not clearly ossifying until 8.2 mm. Principal caudal-fin rays begin to form 1036 Fishery Bulletin 97(4), 1999 Figure 2 Larval Cheilopogon xenopterus. lateral view. (A) 3.5 mm (8710JD, station 3-046); (B) 8.0 mm (8810JD, station 2-049); (C) 8.1 mm (9210JD, station Manta 2). Fin rays shown in dotted lines are based on fins of other specimens of similar size. during embryonic stage X (Table 3) and all 7-t-8 rays present at hatching. All principal rays sup- ported by hypurals through at least 8.2 mm but by 10.5 mm low- ermost principal ray partially sup- ported by haemal spine associated with preural centrum 1. By 13.4 mm lowermost principal ray fully supported and adjacent principal ray partially supported by haemal spine, and by 16.8 mm both rays fully supported by haemal spine. Procurrent caudal-fm rays begin to form shortly after hatching (by 3.8 mm) but full complement of 5- 6 dorsal and 6-8 ventral rays not attained until near end of larval development (by 16.8 mm). Cau- dal fin initially rounded, becomes assymetrical as lower principal rays elongate, beginning at about 5 mm. Pectoral-fin buds form in stage- VIII embryos, almost simulta- neously with beginning of noto- chord flexion, and upper rays form during embryonic stage X. Addi- tion of rays is ventrad, with full complement of 13-15 rays (includ- ing uppermost small, spine-like ray) attained by 9.3 mm (Table 3). Lowermost ray quite small and nearly completely covered by scales in juveniles larger than Table 3 Counts for cleared and stained Cheilopogon xenopterus. In cases where two counts are count from the left side, the second is from the right side. Pectoral fin-ray counts inclu A = anal; Pj = pectoral; P2 = pelvic; C = caudal. given for one category, de the first small, spine the first -like ray value is the D = dorsal; BL(mm) Vertebrae Pleural ribs Epineural inter- muscular Branchiostegal rays Fin rays Abdominal Caudal D A Pi P2 C 3.3 (embryo) 0 0 0 0 2 0 0 3 5 0-(-5-i-5-f0 2.9 0 0 0 0 2 11 10 6 6 0-f7-f8-(-0 3.8 23 2 0 0 6 12 10 13 6 \+l+»+l 5.2 28 15 0 0 7 12 10 11 6 2-f7-i-8+2 6.3 29 14 12 0 7,8 13 10 12 6 •2+7+8+2 8.2 29 14 15 8 9 12 9 12 6 3-i-7-)-8-f3 10.5 29 14 17 14 9 12 11 13 6 3-i-7-(-8-f4 13.4 28 15 20,21 26 9 12 9 15,14 6 4-(-7-i-8-f6 16.8 28 15 22 29 11 13 9 15 6 6-I-7-I-8-1-7 I Watson; Early life history stages of Cheilopogon xenopterus 1037 Cheilopogon xcii (B) juvenile. 36, shown in dotted about 38 mm. Pelvic-fin buds form during embryonic stage IX, at about completion of notochord flexion. All six pelvic-fm rays ap- parently present at hatching. Cartilaginous dorsal- and anal- fin pterygiophores form late dur- ing embryonic development: cleared and stained late stage-X embryo had six dorsal, five anal- fin pterygiophores. Addition and ossification of pterygiophores ap- parently anterior to posterior. Each pterygiophore initially is single cartilage which begins to ossify near middle of lower por- tion (forming proximal radial). Although pterygiophores did not stain well in smaller larvae, at least first three or four anal-fin pterygiophores beginning to os- sify in 3.8-mm specimen and first five or six dorsal- and anal-fin pterygiophores in 5.2-mm speci- men. By 6.3 mm all but posterior three or four dorsal proximal ra- dials ossifying, and distal radials appear to be ossifying adjacent to fin-ray bases. About 8.2-13.4 mm separate ossifications, probably representing middle radials, form on cartilage be- tween proximal and distal radials, but by 16.8 mm each proximal -i- middle radial appears to be a single ossified unit. Dorsal- and anal-fin rays begin to form in stage-Xl embryos; all anal- and most dorsal-fin rays present at hatching. Pigmentation Most elements of basic pattern established during late embryonic development are visible throughout larval period. Principal internal elements of pattern are 1) melanophores under mid- and hindbrain (dif- ficult to see after about 6 mm); 2) row over notochord (difficult to see after about 6 mm); 3) melanophores present around urostyle and proximally on hypurals; 4) melanophores present dorsally, dorsolaterally on gut. Principal external elements are: 1) dorsal mel- anophores covering mid- and hindbrain areas; 2) two somewhat irregular rows along dorsal margin of trunk and tail (increasing to four rows after about 9 mm), and melanophores in each row usually more ^closely spaced along posterior half of dorsal-fin base; 3) two rows along anal-fin base, commonly expanded along posterior half of fin base; 4) patch on each side Figure 3 optenis. lateral view. (A) lan-a. 12. .5 mm (8910M4, station l-OlOl; 4 mm (8910JD. station 1-039). Fin rays (and fin pigment in Al lines are based on fins of other specimens of similar size. over urostyle and hypural area; 5) scattered melano- phores on pectoral, pelvic, and caudal fins. Subsequent additions to internal pigmentation include melanophores spreading around sides of mid- and hindbrain and increasing in number anteriorly over notochord, forming internally on opercular area beginning at about 6-6.5 mm, and spreading vent- rolaterally on gut (primarily anteriorly), surround- ing gut at level of pectoral-fin bases between 12 and 18 mm. Pigmentation on urostyle and hypural area usually increases somewhat. External dorsal pigmen- tation on head slowly increases, spreading forward onto snout after about 13 mm. Melanophores form on upper jaw by about 10 mm. In early juveniles, melanophores over mid- and hindbrain spread ventolaterally, reaching level of pectoral-fin origin by about 36 mm. External melanophores present on central opercular area by about 8 mm (Fig. 2C), form indistinct bar across cheek by about 12-15 mm (Fig. 3A). Barbels initially unpigmented, become densely pigmented distally, usually beginning between 6 and 7 mm. As barbels broaden, melanophores become more concentrated along distal margins and after barbel fusion, connecting membrane becomes in- tensely pigmented along distal margin (by 12-13 mm: 1038 Fishery Bulletin 97(4), 1999 .O^^^ Figure 4 Larval Cheilopogon xenopterus. dorsal vieu- lAi 2.9 mm I8910JD, station 1-015) (B) 10.5 mm I8810JD, station 3-1181; (Ci 13.4 mm (9010JD, station Manta 54). Fig. 4C). Few melanophores usually form along lower jaw after about 7 mm. Soon after hatching, a row of melanophores forms along horizontal septum, originating just behind midtail and spreading cephalad and caudad (prima- rily cephalad) along tail (Fig. 2, A and B). Barred pattern begins to develop at about 6 mm as melano- phores spread upward from pelvic-fin bases and from ventral margins of last three or four preanal myomeres (Fig. 2B). Two more bars begin to spread upward from vicinity of anal-fin rays 5-8 and from midwaj' between pectoral- and pelvic-fin bases (Figs. 2C, 3A) between about 8 and 9 mm. Bars usually do not reach dorsal margin in larvae. Pigmentation on pectoral-fin base sparse until about 8 mm then in- creases, spreads onto adjacent myomeres, and in ju- veniles merges with melanophores spreading down- ward from dorsum to form saddle (Figs. 2C; 3, A and B). Melanophore cluster on hypural area expands to cover almost entire area by about 13 mm. Juveniles thus display six or seven bars (including pectoral saddle); in juveniles >30-40 mm posterior three or four bars most prominent (Fig. 3B). Pectoral-fin pigmentation initially sparse. After hatching, melanophores gradually spread over mem- branes between upper 5-6 rays, a few scattered be- tween all but lowermost 3-4 rays, by about 8 mm (Fig. 2, A-C). Pigment patterns variable in lai-vae Watson: Early life history stages of Cheihpogon xenopterus 1039 <10-12 mm, except that melanophores tend to be more concentrated near distal margin between up- per 5-7 rays and in patch about midway along up- per rays. After about 12 mm concentration midway along upper rays commonly persists, but may dimin- ish or disappear, and distal patch expands to form distinct blotch along upper 6-8 rays, bordered by unpigmented (or sparsely pigmented) margin (Fig. 4C). Distal blotch expands into elongate oval band over much of width of pectoral fin in juveniles (Fig. 3B). Upper 4-6 pectoral rays also may remain sparsely to moderately pigmented, especially proxi- mally and in the blotch midway along upper rays, in larger larvae and juveniles. Apart from upper rays and distal blotch or band, remainder of fin has little pigment after ca. 15 mm. Pelvic-fin pigment may form before hatching; al- ways present after hatching. Sparse initially, com- monly sparse to moderate throughout larval devel- opment (Fig. 4). Pattern variable: melanophores tend to be more concentrated near distal margin of most or all of fin, in distal patch at innermost ray or be- tween two or three middle rays, or in two distal patches at innermost and outermost rays. Occasion- ally melanophores more concentrated distally and mid- way along rays. Distal patch near middle of fin is most common pattern in larger larvae and juveniles. Caudal-fin pigmentation forms in late embryos, initially proximally, mostly or entirely on lower rays. Pigment increases and spreads to about three-quar- ters of length of lower principal rays by ca. 8 mm. Melanophores consistently present on upper part of fin by about 8 mm but pigment here is sparse in lar- vae and sparse to moderate in juveniles. Diagonal band forms on dorsal fin, originating dis- tally between anterior 2-3 rays just after 8 mm (Fig. 2C) and reaching base of rays 6-7 between 18 and 22 mm. Pigmentation may be sparse in lower (=pos- terior) part of band and in larger juveniles band ap- pears more like elongate oval patch along first 5-6 rays. Distal blotch forms at last 2-3 rays beginning at about 12-13 mm. Melanophores form on membranes between (primarily adjacent to) first 5-6 rays after 18 mm. Anal fin unpigmented in larvae, but distal blotch forms at last 2-3 rays in juveniles (Fig. 3B). Comparisons Eggs Most exocoetids have spherical eggs that are about J. 5-3 mm in diameter, lack oil globules, and have filaments on the chorion (Collette et al., 1984). Cheilopogon eggs are typical exocoetids in these char- acters. For most of the Cheilopogon species the fila- ments are evenly distributed as in C. xenopterus. Only C. heteriiriis and C. unicolor have been de- scribed as having a different arrangement, with fila- ments grouped in bipolar clusters (Barnhart, 1932; Miller, 1952; Imai, 1959; Gorbunova and Parin, 1963; Watson, 1996). Among the species with uniform fila- ment distribution, the number of filaments ranges from about 10 to 90; the range in C. xenopterus (about 40-60) apparently is typical (e.g. Collette et al., 1984). Filament length in these species ranges from less than 1 mm to just over 10 mm; most commonly the filaments are long, a character shared with Cypselurus (e.g. Collette et al., 1984). The short filaments of Cheilopogon xenopterus (about 1 mm) are a character shared with Prognichthys (Kovalevskaya, 1982) and with at least some of the members of the C. nigricans species group (Parin and Belyanina, 1996), which includes C. dorsomaculata and C. xenopterus in the eastern tropical Pacific. Eggs have not been described for C. dorsomaculata or the two Prognichthys spe- cies in the eastern tropical Pacific (Parin, 1995) and it is unknown how they might be distinguished from the eggs of C. xenopterus except that late stage Prognichthys embryos should be more densely pig- mented. An unidentified type of exocoetid egg (2.2 mm diameter with ca. 150 short filaments ) occasion- ally co-occurs with C. xenopterus in plankton samples; it might be one of these species (larval Prognichthys are relatively common in the samples). Collette et al. ( 1984) noted that some flyingfishes apparently have a preanal finfold during early de- velopment and suggested that it is an embryonic fea- ture that is lost soon after hatching. There is no prea- nal finfold in C. xenopterus. Larvae and juveniles Larval Cheilopogon range between about 4 and 6 mm (e.g. Collette et al., 1984) and are well developed at hatching. Cheilopogon xenopterus may be a bit smaller than usual, but otherwise is typical of the genus at hatching. Cheilopogon larvae develop a pair of mandibular barbels that persist into the juvenile stage and remain separate or fuse mesially — the membrane joining the barbels either remaining low and simple or becoming broad and fimbriate. Among the Cheilopogon species in the eastern Pacific, C furcatus, C. heterurus hubbsi, and C. papilio retain separate barbels in the juvenile stage. The barbels become fused basally by means of a low, sparsely pig- mented or unpigmented membrane in C. atrisignis and C. spilonotopterus, whereas in C. dorsomaculata and C. xenopterus the basal membrane is somewhat broader and densely pigmented along its margin. 1040 Fishery Bulletin 97(4), 1999 Only one species, C. pinnatibarbatus , is known to have a broad, fimbriate membrane that unites the barbels (Hubbs and Kampa, 1946; Imai, 1959). The barbels become moderately long (Bbl > HL) in C. atrisignis and C. spilonotopterus, whereas the other species retain relatively short barbels (Bbl < HL). The size and condition of the barbels in C. rapanouiensis are unkpown, but in the closely related species C. agoo they are short (Chen, 1987, 1988). Larval pigmentation in Cheilopogon initially ranges from sparse to dense and typically increases gradually during development. The sparsely pig- mented larvae, including C. xenopterus, probably C. atrisignis and C dorsomaculata, and possibly C. spilonotopterus and C. papilio, display a pattern pri- marily of melanophore rows on the dorsum, on the horizontal septum, and on the ventral margin of the tail, and usually bars form in the latter part of the larval stage (Kovalevskaya, 1965, 1977; Chen, 1987. 1988). The more heavily pigmented species, includ- ing C. furcatus, C. heterurus hubbsi, and C. pinnatibarbatus californicus, display a pattern simi- lar to that of larval Cypselurus, with melanophores more evenly distributed over the trunk and at least anteriorly on the tail (sometimes diminishing to dor- sal, midlateral, and ventral rows on the posterior half of the tail) (Hildebrand and Cable, 1930; Hubbs and Kampa, 1946; Watson, 1996). Cheilopogon rapanou- iensis might belong to this latter group as well, if its larvae resemble the moderately pigmented larvae of C. agoo (Chen, 1987, 1988). Juveniles of most spe- cies are barred; about six bars (as in C. xenopterus) is common. Larvae and juveniles of C. xenopterus usually can be distinguished without great difficulty from the other flyingfish species in the eastern Pacific, except perhaps from C. dorsomaculata. The position of the anal-fin origin below dorsal rays 4-7 distinguishes C. xenopterus from Exocoetus, Fodiator, Hirun- dichthys, and Oxyporhamphus, in which the anal- fin origin ranges from just ahead of the dorsal-fin origin to under dorsal rays 1-3, depending on spe- cies (e.g. Imai, 1954; Kovalevskaya, 1964, 1980; Chen, 1988; Watson, 1996). Larger larvae and juveniles of all four genera are further distinguished from C. xenopterus (and all the other Cheilopogon species) by their lack of paired mandibular barbels. Cheilo- pogon xenopterus (and all the other flyingfishes) are also distinguished from Exocoetus by having much more posteriorly placed pelvic fins. Larval Fodiator and Oxyporhamphus are unique in developing a beak beginning at about 6 mm and 8 mm, respectively, whereas Parexocoetus acquires a much smaller beak in the juvenile stage, by about 18 mm (e.g. Collette et al., 1984; Watson, 1996). Cheilopogon xenopterus is rather sparsely pigmented, in contrast to the more general, denser pigmentation of larval Cypselurus, Exocoetus, Fodiator, Parexocoetus, Prognichthys, and the three or four Cheilopogon species noted above (e.g., Hildebrand and Cable, 1930; Imai, 1959, 1960; Kovalevskaya, 1980; Chen, 1988; Watson, 1996). Among the more sparsely pigmented flyingfish lar- vae, Hirundichthys (which may be moderately pig- mented initially, but become more sparsely pig- mented on at least the prepelvic part of the trunk; e.g. Kovalevskaya, 1980) and Oxyporhamphus are easily distinguished from C. xenopterus by the char- acters noted above. In addition, larval Oxypor- hamphus are more elongate, with much shorter pectoral and pelvic fins, and they lack a row of mel- anophores along the horizontal septum before about 9 mm (e.g. Khrapkova-Kovalevskaya, 1963; Watson, 1996) in contrast to before 4 mm in C. xenopterus. Among the five Cheilopogon species in the eastern Pacific with sparsely pigmented (or assumed to be sparsely pigmented I larvae, it usually should be pos- sible to distinguish C. atrisignis and C. papilio from C. xenopterus by a combination of myomere and dor- sal, anal, and (in larger larvae) pectoral-fin ray counts (e.g. Table 1). At larger sizes (small larvae have not been described) larval C. atrisignis and C spilono- topterus can be distinguished from C. .xenopterus by barbel structure and pigmentation, and C. papilio can be distinguished by barbel structure and fin pig- mentation, as noted above. Large larval and juve- nile C. spilonotopterus are more fully and evenly pig- mented on the trunk (by about 14 mm) and lack the barred pattern displayed by larger lai-val and juve- nile C. xenopterus (e.g. Kovalevskaya, 1977; Chen, 1987, 1988). Cheilopogon papilio likewise become more generally pigmented than C. xenopterus, al- though four or five bars remain visible through at least 21 mm. Juvenile C. dorsomaculata smaller than about 40 mm bear a striking resemblance to C .xenopterus but can be distinguished by small differ- ences in meristic characters (Table 1) and have more sparsely pigmented (nearly unpigmented) pelvic fins than do C. .xenopterus. Larvae of C. dorsomaculata have not been described, but the similarity of the small ju- veniles to C. .xenopterus suggests that it may be diffi- cult to distinguish the larvae of the two species. Acknowledgments I thank the scientific and ship crews of the RV David Starr Jordan and RV McArthur, and especially Bob Pitman, Steve Reilly, and Tim Gerrodette of the SWFSC Marine Mammals Division, for collecting and providing access to the plankton samples containing Watson: Early life history stages of Cheilopogon xenopterus 1041 the specimens that made this study possible. Lucy Dunn and Jeanne Haddox sorted the eggs and lar- vae from the plankton samples. Student interns Hugh Bang and Patty Lopez aided in taking the ra- diographs. Geoff Moser and Bruce Collette read the manuscript and made helpful suggestions. Literature cited Barnhart, P. S. 1932. Notes on the habits, eggs, and young of some fishes of southern California. Bull.Scripps Inst. Oceanogi-.. Tech. Ser. 3:87-99. Belyanina, T. N. 1993. Early stages of development of the east Australian flying fishes (family Exocoetidaei. Trudy Inst. Okeanol. Akad. Nauk. SSSR, 128:108-146. [In Russian.) Bleeker, P. 1866. Sur les especes d"Exocet de I'lnde Archipelagique. Neder. Tijdschr. Dierk. 3: 10.5-129. Breder, C. M., Jr. 1938. A contribution to the life histories of Atlantic Ocean flyingfishes. Bull. Bingham Oceanogr. Collect. Yale Univ. 6(51:1-126. Brown, D. M., and L. Cheng. 1981. New net for sampling the ocean surface. Mar. Ecol. Prog. Ser. 5:225-227. Chen, C.-H. 1987. Studies of the early life history of flying fishes ( fam- ily Exocoetidae) in the northwestern Pacific. Taiwan Mus. Sp. Publ. Ser. 7:1-203. (In Chinese.] 1988. Beloniformes. In M. Okiyama led.). An atlas of the early stage fishes in Japan, p. 259-301. Tokai Univ. Press. Tokyo, 1154 p. [In -Japanese.] Clarke, H. W. 1936. New and noteworthy fishes. The Templeton Crocker expedition of the California Academy of Sciences. 1932. Proc. Calif Acad. Sci. Ser 4. 21(29):383-396. Collette, B. B., G. E. McGowen, N. V. Parin, and S. Mito. 1984. Beloniformes: development and relationships. /;i H. G. Moser, W. J. Richards. D. M. Cohen, M. R Fahay. A. W. Kendall Jr, and S. L. Richardson (eds. I, Ontogeny and sys- tematics of fishes, p. 335-354. Am. Soc. Ichthyol. Herpetol. Spec. Publ. 1, 760 p. Dasilao, J. C, Jr., K. Sasaki, and O. Okamura. 1996. The hemiramphid. Oxyporhamphini. is a flyingfish (Exocoetidaei. Ichthyol. Res. 44(2>:101-107. Fowler. H. W. 1944. Results of the fifth George Vanderbilt expedition ( 19411 (Bahamas, Caribbean Sea. Panama, Galapagos Ar- chipelago and Mexican Pacific islands I. The fishes. Monogr Acad. Nat. Sci. Philad. 6:57-529. Gibbs, R. H., Jr., and J. C. Staiger. 1970. Eastern tropical Atlantic flyingfishes of the genus Cypselurus (Exocoetidae). Stud.Trop. Oceanogr (Miami) 4(2):432-466. Gilbert, C. H. 1890. A preliminary report on the fishes collected by the steamer Albatross on the Pacific coast of North America during the year 1889, with descriptions of twelve new gen- era and ninety-two new species. Proc. U. S. Nat. Mus. 13 (797):49-126. Gillett, R., and J. lanelli. 1991. Aspects of the biology and fisheries of flyingfish in the Pacific islands. FAOAJNDP Reg. Fish. Support Prog., Field Doc. 91/7, 35 p. Gorbunova, N. N., and N. V. Parin. 1963. Development of the flying fish Cheilopogon . jP-^" Figure 1 Location of Alaska villages where walrus harvest information is collected through the Walrus Harvest Monitor Project (O) and the Marking Tagging and Reporting Program (•:. Note that tusks may be tagged in selected locations outside the walrus' range. Owing to funding and logistical constraints that prevent the year-round, statewide implementation of the WHMP and imperfect hunter compliance with the MTRP, not all harvested walruses are recorded. The FWS must therefore rely on analytical methods to estimate the size of the total annual harvest based on the best available information. Here we describe a new method for estimating the size of the annual walrus harvest in Alaska using a correction factor for noncompliance with the MTRP rule. Materials and methods Prior to the initiation of the MTRP, the statewide walrus harvest was approximated by using a predic- tion equation applied to WHMP data. The predic- tion equation was based upon the historic relation- ship of the size of the spring harvest at WHMP-moni- tored villages in relation to the remaining hunting villages in Alaska. This information was obtained from the Alaska Department of Fish and Game, which administered a statewide harvest monitoring program between the years of 1960 and 1978. To es- timate the statewide harvest, the FWS used a least- squares regression to describe the relation between the Gambell, Savoonga, and Diomede harvests to the statewide totals. The resulting prediction equation (Equation 1, /•2=0.239) was subsequently used to es- timate the total statewide harvest on the basis of data collected annually after 1978 through the WHMP at Gambell, Savoonga, and Diomede. A^ = 459.7 + 1.26 ( WHMP7-„,„,), (2) where WHMP Total N = total statewide harvest estimate; and - total number of retrieved walruses re- corded during monitored spring hunts at select villages (Diomede, Gambell, and Savo- onga) for a given year. The evolution of the MTRP as a year-round, state- wide monitoring program provided the opportunity to improve the reliability of harvest estimates by using current data. Because there is spatial and tem- poral overlap between the two programs, the WHMP can serve as a baseline for evaluating compliance with the MTRP program. A correction factor for non- compliance can be estimated as a proportion by us- ing the following equation: R WHMP, Total MTRP,y„!^p (2) i NOTE Garlich-Miller and Burn; Estimating the harvest of Odobenus rosmarus divergens in Alaska 1045 Table 1 Walrus harvest data fr om the Wah'us Harvest Monitor Project (WHMP) and Mark ing, Tagging, and Reporting Program (MTRP), | 1992-1997. Correction Standard factor error Prediction Proportional 95'"/( CI of Year V^tiMPr,„„, MTRP,,„,„ (R) SE(R) MTRPr..,,, Equation 1 Equation 3 Equation 3 1992 820 976 1700 1491' 1993 710 708 1.003 0.21 1189 1353 1192 703-1682 1994 971 799 1.215 0.17 1328 1681 1614 1171-2056 1995 1204 785 1.534 0.17 1096 1974 1681 1316-2046 1996 1220 773 1.578 0.33 1572 1994 2481 1464-3497 1997 856 574 1.491 0.09 1045 1537 1558 1373-1742 ' Ttieca culated estimate was less than the va lue ofA/TffP.,;,,, ,. The A/rflP.,,„„ was therefore used as a minimum estimate of the statewide walrus harves t. where R MTRP WHMP - correction factor for noncompliance; and a subset of the numljer of wahaises recorded through the MTRP with re- ported kill dates and locations that coincide with WHMP operations (pre- sumed to have been recorded through both the MTRP and the WHMP). The difference between the two methods was not sta- tistically significant for any year. Comparisons of the 1992 MTRP and WHMP harvest data suggests that the WHMP monitors were unable to account for all the retrieved walruses landed during the monitoring period. In this instance, the MTRPj^^^^^^i was considered the minimum estimate of the statewide harvest. The standard error of this ratio estimate is calcu- lated according to Snedecor and Cochran ( 1967, p. 537) by treating each of the three villages as inde- pendent estimates of overall compliance. Data from the village of Wales were not used in the analysis because of small harvests and a lack of independence between the WHMP and MTRP programs in that village. Assuming the estimated compliance with the MTRP is uniform throughout the full range of wal- rus hunting villages, the total MTRP harvest can be adjusted as follows: TV = MTRProtai x R, (3) where MTRP Total ■ the total number of walrus re- corded through the MTRP. Results Data from the MTRP and WHMP are currently avail- able for the years 1992-97 (Table 1 ). Estimated com- pliance with the MTRP (the reciprocal of R) ranged from 63% to 99*7^. Harvest estimates based on the prediction equation and the proportional equation differed depending on the proportion of animals re- corded through each of the two monitoring programs. Discussion Prior to 1997, the FWS used WHMP data collected annually at Gambell, Savoonga, and Diomede as an index for estimating the total statewide harvest of wal- rus. One drawback of this method has been that the equation assumes that the relation between the hunt- ing success of WHMP and non-WHMP villages is con- stant over time (Fay et al., 1997). hi fact, recent har- vest data have shown that the relative hunting suc- cess of each village is highly variable. The annual har- vest at each village is subject to large interannual varia- tion, presumably as a result of weather and ice pat- terns affecting the availability of walruses to hunters at a given geographical location (Garlich-Miller-^). One advantage of the proportional equation (Eq. 3) over the prediction equation (Eq. 1) is that it can account for variability in the relative success of hunt- ing villages. For example, on the basis of comparable numbers of walrus recorded through the WHMP in Gambell, Savoonga, and Diomede in 1995 and 1996, the prediction equation produced remarkably simi- lar harvest estimates for the two years (1974 and 1994 walrus, respectively). Although MTRP compli- ance rates were also comparable for these years, the proportional equation, in which statewide MTRP data wei'e used, produced hai-vest estimates that were mark- edly different ( 1681 and 2481 walrus, respectively). We 1046 Fishery Bulletin 97(4), 1999 believe that this difference, which reflects inter- annual variability in the hunting success of non- WHMP villages, more accurately reflects statewide harvest levels for these two yeai's. For the reasons stated above, the FWS has adopted the use of the proportional equation to estimate the size of the walrus harvest. In 1992, the assumption that all retrieved walrus were redsorded during WHMP operations was not met. In this instance we considered the uncorrected MTRP data as a minimum harvest estimate for that year. The WHMP was re-initiated in 1992 after a two-year hiatus; the lack of experienced personnel likely con- tributed to the poor monitoring results that season. Since that time, program managers have attempted to improve the program by hiring experienced moni- tors and additional village assistants to meet all re- turning boats and by emphasizing the importance of recording all retrieved animals (Dickerson-^). Another assumption of the proportional equation is that compliance in the villages of Gambell. Savoonga, and Diomede is representative of non- WHMP villages. Compliance was variable between villages and years, suggesting that thei'e was little correlation between the two progi'ams. Since conduct- ing analyses of MTRP compliance, information and education efforts in these villages appear to have been effective at increasing compliance. The accu- racy of this method could be improved by increasing the number of villages in the WHMP. In both the prediction and proportional equations, estimates are based on the number of walrus that are retrieved by hunters. Fay et al. ( 1994) estimated that 427c of walrus struck by hunters are not re- trieved and subsequently die at sea. In order to esti- mate total human-caused removals from the Pacific walrus population, harvest estimates are adjusted to account for animals struck and lost. Accurate harvest data are vital to the management of the Pacific walrus population. Harvest data are incorporated into stock assessment reports that chart the status and trend of the population. The stock assessment process compares estimates of human- caused mortality with a calculated potential biologi- cal removal (PBR) level to determine the status of a stock. One of the reasons a stock may be designated as "strategic" depends upon whether or not its level of human-caused mortality exceeds the calculated PBR level (Wade, 1998). Since 1992, most human caused mortality affecting the Pacific walrus popu- lation has been associated with walrus hunting ac- tivities in Alaska and Chukotka. Between 1992 and 1996 the combined annual take of walrus in the U.S. and Russia averaged 4869 walrus per year (Gorbics et al.''). Russian harvest data are currently unavail- able for 1997. Because annual estimates of human caused mortality have been lower than the calculated PBR of 7533 the population has been classified as non- strategic (Gorbics et al.''). It is essential that harvest monitoring in both nations be maintained in order to accurately assess the impact of the harvest to this stock. In summary, this new method of harvest estima- tion uses data from both harvest monitoring pro- grams to account for interannual and intervillage variability in hunting success and applies a correc- tion factor to adjust total harvest estimate to account for noncompliance with the MTRP. The accuracy of harvest estimates is therefore dependent upon the degi'ee of hunter compliance with the MTRP rule. It is hoped that through ongoing information and edu- cation efforts that explain the importance of accu- rate harvest data, the understanding of and compli- ance with monitoring programs will improve. Acknowledgments We would like to acknowledge the hard work and dedication of the MTRP taggers and WHMP harvest monitors. Harvest monitoring activities were coor- dinated by Larry Dickerson, Polly Hessing, Dana Seagars, and Wells Stephensen. This manuscript was improved by constructive comments offered by the fol- lowing reviewers: Susan Lapkass, Larry Dickerson, Susan Hills, and Mark Udevitz. Literature cited Burn, D. M. 1998. Estimation of hunter compliance with the Marine Mammal Marking. Tagging, and Reporting Program for walru.s. Wild. .Soc. Bull. 26i 1 1:68-75. Fay, F. H., J. J. Burns, S. W. Stoker, and J. S. Grundy. 1994. The struck-and-lost factor in .Ma.skan walrus harvests. Arctic 47(4 ):.368-373. Fay, F. H., L. L. Eberhardt, B. P. Kelly, J. J. Burns, and L. T. Quakenbush. 1997. Status of the Pacific walrus population, 19.50-1989. .Mar. Mamm. Sci. 13:537-565. Ray. D. J. 1975. The eskimos of the Bering Strait, 1650-1898. Univ. Washington Press. Seattle, WA, 305 p, Snedecor. G. W., and W. G. Cochran. 1967. .Statistical methods, sixth ed. The low-a State Univ. Press, Ames, lA, 593 p. Wade, P. R. 1998. Calculating limits to the allowable human-caused mortality of cetaceans and pinnipeds. Mar. Mamm. Sci. 14(ll;l-37. Gorbics, C. S.. J. L. Garlich-Miller, and S. L. Schliebe. 1997. Draft Alaska marine mammal stock assessments 1997: sea otter, polar bear and walrus. U.S. Fi.sh and Wildlife Service, Marine Mammals Management, Anchorage, AK. Admin. Report, 129 p. ^ t i: 11 1' ■ a « IT I' 104 7 The trophic role of Atka mackerel, Pleurogrammus monopterygius, In the Aleutian islands area Mei-Sun Yang Alaska Fisheries Science Center 7600 Sand Point Way NE Seattle, Washington 98115-0070 E-mail address Mei Sun Yangia noaa gov Atka mackerel (Pleurogrammus monopterygius) is presently the most abundant marine fish species in the Aleutian Islands. Its exploit- able biomass in 1998 was estimated to be 535,500 metric tons (t) (Lowe and FritzM. The commercial catch of Atka mackerel increased from 26,740 1 in 1991 to 104,000 1 in 1996 (Lowe and FritzM. Atka mackerel has been identified as food for other marine fishes, i.e. Pacific cod (Ga- dus maci-ocephalus) (Andriyashev, 1937), coho salmon (Oncorhynchus kisutch ) (Wing, 1985; Volkov et al., 1995; Davis-), arrowtooth fiounder (Atheresthes stomias) (Mito, 1974); sea birds such as thick-billed murres (Uria lomvia) (Ogi, 1980); and marine mammals such as Steller sea lions (Eumetopias jubatus) (Merrick et al., 1997). The western stock of Steller sea lions has recently been listed as endangered. A connec- tion between the observed decline in Steller sea lions and the sea lion's dependence on commercially ex- ploited fish species such as Atka mackerel as prey has been sug- gested (Merrick et al., 1997). Be- cause of the great abundance of Atka mackerel and its important role as food for marine fish and marine mammals, it is essential to understand the trophic role of Atka mackerel in the Aleutian Islands area. The objective of this study was to describe the role of Atka mackerel as prey and as predator in the Aleutian Islands ecosystem. Methods Study area From July to September 1991, the Resource Assessment and Conser- vation Engineering (RACE) Divi- sion at the Alaska Fisheries Science Center (AFSC), Seattle, Washing- ton, conducted a groundfish re- sources trawl survey in the Aleu- tian Islands region. This area (Fig. 1 ) was divided into four major geo- graphical units: southeastern Bering Sea area (from long. 165°W to 170°W, 42,604 km2); eastern Aleutian area (long. 170"W to 177°W, 338,818 km^); central Aleu- tian area (long. 177°W to 177°E, 322,234 km-); and western Aleu- tian area (long. 177°E to 170°E, an area of 343,083 km2). Two char- tered fishing vessels. Ocean Hope and Green Hope, were used. Both vessels used standard RACE Division Poly-Nor'eastern, hard bottom, high-opening bottom trawls constructed of 12.7-cm (5- inch) stretched-mesh polyethylene web, with a 3.2-cm (1-1/4 inch) stretched-mesh nylon liner in the codend to retain smaller specimens (Harrison, 1993). Trawls were towed at 3 knots for 30 minutes. Scientists from the Resource Ecol- ogy and Fisheries Management (REFM) Division, Resource Ecology and Ecosystem Modeling Task, col- lected fish stomach samples during this survey. Sample collection Fish stomach samples were col- lected during trawling operations by scientists aboard the charter vessel Green Hope. Both Atka mackerel and their potential fish predators were sampled and mea- sured, and stomach samples were collected. Because of the previously demonstrated high similarity ( >QQ'^/c ) in diet between arrowtooth flounder and Kamchatka flounder (Yang and Livingston, 1986), no stomach samples of Kamchatka flounder were collected. Before excising a stomach, fish were examined for evidence of regurgitation or net feeding. If a fish had food in its mouth or around the gills, or if its stomach was everted or flaccid, the fish was categorized as having re- gurgitated food and the specimen was discarded. If a predator had fresh food (usually fish) sticking out of its mouth or throat, it was catego- rized as a net-feeding fish and was also discarded. When a sampled stomach was retained, it was put into a cloth stomach bag. A field tag with the species name, fork length (FL), and haul data (vessel, cruise, haul number, and specimen number) was also put into the bag. All of the samples collected were then pre- served in buckets containing a 10% buffered formalin solution. When ' Lowe. S. A., and L.W. Fritz. 1997. Atka mackerel. In Stock assessment and fish- ery evaluation report for the groundfish resources of the Bering Sea/Aleutian Is- lands regions as projected for November 1997, compiled by the plan team for the groundfish fisheries of the Bering Sea and Aleutian Islands. North Pacific Fishery Management Council, P.O. Box 103136, Anchorage, AK 99.510. - Davis, N. D. 1990. U.S. -Japan coopera- tive high seas salmonid research in 1990: summary of research aboard the Japanese research vessel Hokuho marii, 4 June to 19 July. (INPFC Doc. 3.508) FRI-UW- 9010. Fish. Res. Inst.. Univ. Washington, Seattle, WA 98195, 24 p. Manuscript accepted 11 December 1998. Fish. Bull. 97(41:1047-1057 11999). 1048 Fishery Bulletin 97(4), 1999 Study area Alaska S5N Western Central Eastern ■*»^ie-*" M Figure 1 The subareas of the Aleutian Islands trawl survey area. the samples arrived at the laboratory, they were transferred to 70*^ ethanol before the contents were analyzed. Stomach content analysis In the laboratory, stomach contents were first blot- ted with a paper towel and the wet weight recorded to the nearest one-tenth of a gram. After obtaining the total weight of the contents, these were placed in a petri dish and examined under a dissecting mi- croscope. Each prey item was classified to the lowest practical taxonomic level. The numbers of noncom- mercially important prey were not counted; instead the percent total volume of these prey items in stom- achs was visually estimated. Prey weights and num- bers of commercially important crabs and fish were recorded. If walleye pollock otoliths were found, otolith lengths were measured and the pollock's stan- dard length (SL) was derived through an otolith length-fish length regression table (from REFM's Ageing Task ). Other fish otoliths were not treated as walleye pollock otoliths either because they were not found as frequently as walleye pollock or were diffi- cult to identify to the species level. Standard lengths of prey fish, carapace widths (CW) of Tanner crabs (Chionoecetes spp.), and Korean horse-hair crabs (Erimacrus isenbeckii) were also recorded. Data analysis Atka mackerel as prey and predator The general diet of each species was analyzed to determine the overall percent total weight that each prey item rep- resented in the stomach and to identify predators of Atka mackerel. Marine fishes that were mainly fish eaters were then selected for an analysis of popula- tion consumption of Atka mackerel. Stomach content analysis of Atka mackerel were performed to under- stand the role of Atka mackerel as a predator in the Aleutian Islands area. Population level consumption of Atka mackerel by marine fish Following their identification as preda- tors, the consumption of Atka mackerel by four ma- rine fish species. Pacific cod, Pacific halibut, arrowtooth flounder, and Kamchatka flounder, was estimated. The amount of Atka mackerel consumed NOTE Yang: The trophic role of Pleurogrammus monopteiygius 1049 at the population level was calculated by using the following equation: ^C, =DR, xDxB, xP,, where C, DR. D = B. the consumption (by weight) of Atka mackerel by size group / of a predator species; the daily ration (as a proportion of body weight per day) of predator size group / (Table 1); the number of days when Atka mack- erel were vulnerable to predation ( from 1 May to 30 September in this study); the biomass of the predator size group i (Table 2); and the proportion by weight of prey Atka mackerel in the diet of predator size group / (Table 3). Daily rations were calculated by using annual growth increments and conversion efficiencies (Brett and Groves, 1979). Predator biomass data were ob- tained from the RACE Division database at the AFSC. The percent by weight of prey Atka mackerel in the diet of predators (P^) was calculated for arrowtooth flounder, Pacific cod, and Pacific halibut. Because of the high similarity (>60'^) between the diets of Kamchatka flounder and arrowtooth floun- der (Yang and Livingston, 19861, the percent by weight of Atka mackerel that arrowtooth flounder consumed was also used for Kamchatka fiounder. Results Atka mackerel as prey of marine fishes Eleven commercially important groundfish species were collected: Pacific cod, Gadus macrocephalufi: walleye pollock, Theragra chalcogramma: Pacific halibut, Hippoglossus stenolepis; arrowtooth fioun- der, Atheresthes stomias; Greenland turbot, Rein- hardtius hippoglossoides; Atka mackerel, Pleuro- grammus monopterygius; Pacific ocean perch, Sebastes alufus; northern rockfish, S. aleutianus; shortraker rockfish, S. borealis; and shoi'tspine thornyhead, Sebastolobus alascauus. According to stomach content data, Atka mackerel (mostly juve- niles) were found in the stomachs of Pacific cod. Pa- cific halibut, and arrowtooth flounder. The detailed food habits of these species were previously reported (Yang, 1996). The mean percent weights of Atka mackerel consumed by the predators are given in Table 3 by predator size and subarea. Arrowtooth Table 1 Daily ration (D/-, , in terms of proportion of body weight per day) of four groundfish predators in the Aleutian | Islands area in 1991. Predator size Daily Predator species (cm) ration Arrowtooth flounder and <20 0.009 Kamchatka flounder 20-39 0.005 >40 0.008 Pacific cod <30 0.011 30-59 0.011 >60 0.008 Pacific halibut 50-79 0.004 >80 0.007 flounder >40 cm FL consumed large amounts (36.6% by weight) of Atka mackerel in the western Aleutian area, whereas Pacific cod >60 cm FL and Pacific hali- but (especially fish >80 cm FL) both consumed great amounts of Atka mackerel in the central Aleutian area (42'^/i and 72.9% by weight, respectively). Atka mackerel as predator A total of 238 Atka mackerel stomachs were analyzed, 231 (97%) of which contained food. Atka mackerel ranged from 22 to 44 cm FL, with a mean and SD of 33.5 cm and ±4.5 cm, respectively. Table 4 lists the percent by weight of the prey found in Atka mack- erel stomachs. More than 90% of the total weight of stomach contents was invertebrates and less than 5% was fish. Euphausiids (mainly Thysanoessa inei'jnis) were the most important prey, represent- ing more than 28''^ of the total stomach contents. Calanoid copepods were another important prey of Atka mackerel, representing approximately 23% of the total stomach content weight. Larvaceans and Themisto sp., a hyperiid amphipod, occurred fre- quently but each represented less than 10% of the total weight of stomach contents. Squid were another invertebrate prey of Atka mackerel; they represented 5% of the total weight of stomach contents. Several fish species were preyed on by Atka mack- erel. Walleye pollock represented 2.4% by weight of the total stomach contents. Myctophids, bathylagids, zoarcids, stichaeids, and pleuronectids were also found. However, each of them i-epresented less than V'f of the total weight of stomach contents. Canni- balism of eggs was also observed in Atka mackerel stomachs, and eggs represented 5.5% of the total weight of stomach contents. Figure 2 shows the percentage by weight of the main prey items for different Atka mackerel size 1050 Fishery Bulletin 97(4), 1999 Table 2 Biomass (B, ; in tons of predators of Atka mackerel by size groups in three subareas (W, C E) around the Aleuti an Islands and southeastern Bering Sea (SBS) in 1991. ATF = arrowtooth flounder; PCOD = Pacific cod; PH = Pacific halibut. ATKA = Atka mackerel; KF = Kamchatka flounder; W = western Aleutian; C = central Aleutian; E = eastern Aleutian. Species (cm) W C E SBS Total ATF - <20 1.9 0.6 29.4 37.6 20-39 947.8 728.1 2162.1 4276.3 >40 3796.2 1571.1 4733.5 8315.8 Total 4745.9 2299.8 6925.0 12,629.8 26.601 PCOD <30 12.5 153.7 138.6 151.6 30-59 5507.2 9791.5 15,010.9 3886.2 >60 57,847.8 28.355.0 46,159.4 7448.3 Total 63,367.6 38,300.2 61,308.8 11,486.1 174,463 PH <50 0 157.5 204.5 1857.8 50-79 560.1 1795.1 6813.4 2891.8 >80 4745.9 5267.2 11,123.4 4207.9 Total .5306.0 7219.8 18,141.3 8957.7 39.625 ATKA <25 2537.9 2267.6 2000.0 0 26-35 142,607.1 86,328.5 65,407.6 0 >35 175,047.2 185,358.4 2629.7 0 Total 320,192.3 273,954.5 70,037.4 0 664,184 KF <20 0 0 9.3 0 20-39 280.9 251 442.2 4.8 >40 5281.5 9028.7 1114.6 881.7 Total 5562.4 9279.7 1566.1 886.6 17.295 ! Table 3 Mean percent weight (P,> of Atka mackerel cons umed by marine fish in three subareas (W. C, 5) around the Aleu- tian I slands and southeastern Bering Sea (SBS m 1991. ( — ) indicates no sample. Values in parenthe ses are sample sizes. ATF = arrowtooth flounder; PCOD = Pacific cod; PH = Pacific h alibut. W = western Aleutian; C = central Aleutian; E = eastern Aleutian. Species (cm) W C E SBS ATF <20 — 0 (1) 0 (6) 20-39 0 (20) 0 (9) 0 (56) 0(36) >40 36.6 (19) 0 (3) 0 (28) 14.3 (16) Total 32.1 (39> 0(12) 0 (85) 7.1 (.58) PCOD <30 — — 0 i3i 0(31) 30-59 5.0 (78) 14.5(44) 0 (68) 0(28) >60 25.3(117) 41.7(23) 5.6(145) 0 (14) Total 24.1(195) 40.2(67) 3.6(316) 0(73) PH <.50 0 (3) 0 (6i 0 (5) 0(47) 50-79 4.0 (10) 36.0 (7) 0 (39) 9.8(33) >80 38.5 (9i 72.9 (5) 0.2 (10) 0 (6) Total 28.2 (22) 33.8(18) 0.1 (54) 4.0(86) groups. The most important food of the smallest (<25 cm) size group of Atka mackerel were calanoid copepods ( 35^f by weight ) and larvaceans ( SS'/r ). The most important prey of Atka mackerel between 26 and 35 cm FL were calanoid copepods (35%) and eu- phausiids (29% ). The largest size group (>35 cm) of Atka mackerel consumed a substantial amount of calanoid copepods ( 18% ), in addition to euphausiids (22%). Miscellaneous fish (walleye pollock, cottids, myctophids, and zoarcids) represented approximately 9% of the total weight of stomach contents of the larg- est size group of Atka mackerel. Atka mackerel eggs (as evidence of cannibalism) were found almost ex- clusively in the largest size group, representing 11% of total weight of stomach contents. Walleye pollock were the most common fish spe- cies in the Atka mackerel diet. The size of walleye pollock consumed ranged from 49 to 63 mm SL with a mean and SD of 56.2 mm and ±4.8 mm, respec- tively. The cottids consumed by Atka mackerel ranged from 12 to 21 mm SL with mean and SD of 15.9 mm and ±3.1 mm, respectively. In addition, Atka mack- erel stomach samples contained one bathylagid NOTE Yang: The trophic role of P/eurogrammus monopterygius 1051 Table 4 Mean percent total weight C/rW) and mean percent of frequency of occurrence C/rFO) of the prey items of Atka mackerel iPleurogrammus monoptgerygius) collected in the Aleutian Islands area in 1991. Prey name 9cW %FO Prey name %W %F0 Polychaeta (worm) 0.62 2.59 Caridea (shrimp unidentified) 0.11 15.45 Gastropoda (snail) 0.88 34.86 Hippolytidae (shrimp) 0 0.48 Pteropoda 0.03 3.89 Pandalidae (shrimp) 0.12 0.32 Cephalopoda (squid and octopus) 0.07 5.29 Paguridae (hermit crab) 0 2.06 Teuthoidea (squid) 4.67 27.26 Majidae (spider crab) 0.06 3.81 Octopoda (octopus) 0.09 1.27 Hyas sp. (lyre crab) 0.02 3.66 Crustacea (unidentified) 4.89 6.98 Chionoecetes sp. (snow and Tanner crab) 0.06 6.72 Ostracoda ( unidentified 1 0.04 12.79 Erimacrus isenbeckii (Korean hoi-se-haii" crab ) 0 0.32 Calanoida (copepod unidentified) 22.68 61.12 Cancridae (crab) 0.03 0.32 Eucalanidae ( copepod 1 0.23 5.49 Ectoprocta (bryozoan) 0.02 1.98 Eucalanus sp. (copepod) 0.01 1.29 Chaetognatha (arrow worm) 0.79 23.71 Pseudocatanus sp. (copepod) 0 0.34 Lar\'acea Copelata 9.05 72.28 Candacia columbiae (copepodi 0.51 45.00 Chondrichthyes 0.01 1.66 Mysidacea (mysid unidentified) 0.01 1.59 Osteichthyes Teleostei (bony fish unidentified) 0.1 1.27 Mysidacea Mysida i mysid ) 0.01 0.63 Bathylagidae (deepsea smelts) 0 0.48 Acanthomysis pseudomacropsis (mysid) 0 0.48 Myctophidae (lanternfish) 0.75 0.95 Cumacea (cumacean unidentified) 0.02 5.29 Theragra chalcogramma (walleye pollock) 2.4 2.06 Amphipoda (amphipod unidentified) 0.02 1.82 Zoarcidae (eelpout) 0.67 0.32 Gammaridea (amphipod) 0.84 16.36 Atka mackerel eggs 5.52 6.67 Themisto sp. (amphipod) 2.41 59.16 Stichaeidae (prickleback) 0.01 0.32 Caprellidea (amphipod) 0.02 2.45 Pleuronectidae (flatfish) 0.01 0.32 Euphausiacea (euphausiid unidentified) 27.46 60.08 Unidentified organic material 10.39 28.49 Euphausia sp. (euphausiid) 0.03 0.48 Fishery discards 0.58 0.32 Euphausia pacifica (euphausiid) 0.02 0.48 Thysanoessa mermis (euphausiid) 0.75 0.48 Total prey weight 700.55 g Thysanoessa inspinata (euphausiid) 0.14 0.32 Total nonempty stomachs 231 Thysanoessa longipes (euphausiid) 0.1 0.48 Total empty stomachs 7 Thysanoessa spinifera (euphausiid) 0.19 0.95 Total hauls 21 (11 mm SL), one zoarcid (61.4 mm SL), one myctophiii (59 mm SL), anci one stichaeid (29 mm SL). Consumption of Atka mackerel at the population level The dominant predator species. Pacific cod and Pacific halibut, consumed large amounts of Atka mackerel (Table 5). The estimated amount of Atka mackerel con- sumed by each species varied by predator size as well as by the location where the predator was collected. For Pacific cod, large fish (>60 cm) in the western and cen- tral Aleutian areas consumed between 14,000 and 18,000 t of Atka mackerel. For Pacific halibut, large fish (>80 cm) con- sumed between 2000 and 4000 t in the western and central Aleutian Islands ar- eas. Consumption of Atka mackerel by arrowtooth flounder was found in samples n=11 132 CL 100 I •"■n- Axv^>^xy Prey Key : Misc. prey Cephafopod 1 Other crustacean SCalanoid 'Euphausiid ALarvacean jMisc. fish ■ Atka mackerel eggs 20 Pre(Jator fork length (cm) Figure 2 Variation in the main food items of Atka mackerel, by predator size, in the Aleutian Islands area in 1991. n = sample size. 1052 Fishery Bulletin 97(4), 1999 Table 5 Biomass (t) of Atka mackerel population consumed by marine fishes in three subareas (W,C E) around the Aleutian Islands and the southeastern Bering Sea (SBS) area in 1991. (— ) indicates no sample. Values in parentheses are % total b omass. ATF = arrowtooth flounder; PCOD = Pacific cod; PH = Pacific halibut; KF = Kamchatka fio ander. W = western Aleutian; C = central | Aleutian ; E = eastern Aleut an. Species cm) W C E SBS Total ATF " <20 — — 0 0 0 20-39 0 0 0 0 0 >40 1702 (3.5) 0 0 1455(3.0) 3157 (6.5) Total 1702 (3.5) 0 0 1455 (3.0) 3157 (6.5) PCOD <30 — — 0 0 0 30-59 460 (1.0) 2394 (4.9) 0 0 2854 (5.9) >60 17,914(36.9) 14.487(29.9) 3170(6.5) 0 35,570(73.3) Total 18,374(37.9) 16,881 (34.8) 3170(6.5) 0 38,424(79.2) PH <50 0 0 0 0 0 50-79 14 (0.1) 395 (0.8) 0 173 (0.4) 582 (1.2) >80 1958 (4.0) 4114 (8.5) 19(0.1) 0 6091(12.6) Total 1972 (4.1) 4509 (9.3) 19(0.1) 173(0.41 6673(13.8) KF <20 — — 0 0 0 20-39 0 0 0 0 0 >40 126 (0.3) 0 0 154(0.3) 280 (0.6) Total 126 (0.3) 0 0 154(0.3) 280 (0.6) All species 22,173(45.6) 21,390(44.1) 3189(6.6) 1782(3.7) 48,534 from the western Aleutian and southeastern Bering Sea areas. Because the diet of arrowtooth flounder and Kamchatka flounder are very similar (Yang and Livingston, 1986), the daily ration values (Table 1) and the values of the mean percent of the Atka mack- erel consumed by arrowtooth flounder (Table 3) were used for the calculation of the consumption of Atka mackerel by Kamchatka flounder. The total biomass of Atka mackerel consumed by the main marine fishes in the Aleutian Islands areas was 48,534 t (Table 5). The number of Atka mackerel consumed by ma- rine fishes was also estimated from the biomass con- sumed (Table 5) and the prey length recorded from stomach content analysis. The total number of Atka mackerel consumed by marine fishes in the Aleutian Islands area was about 284 million (Table 6), 84% of which were consumed by Pacific cod. The size and age composition of Atka mackerel consumed by groundfishes is listed in Table 7. Age-2 fish repre- sented about 40'7f of the total numbers of Atka mack- erel consumed by groundfishes, including predators of all sizes from different areas. Age-1 and age-3 fish accounted for 30% and 22% of the total numbers con- sumed, respectively. Age-0 Atka mackerel (4%f I were consumed only by Pacific cod >60 cm FL in the cen- tral Aleutian area. Some age-4 and age-6 Atka mack- erel were also found in marine fish stomachs, but they accounted for only about 4% of the total num- ber consumed compared with about 25% of the Atka mackerel that were age 3. In terms of biomass, the age 2-1- Atka mackerel consumed by marine fishes (41,071 1) represented about 3.2% of the estimated Atka mack- erel population biomass (age 2-i-) (Lowe and Fritz^). Discussion Atka mackerel as prey of marine fishes In addition to groundfish, such as Pacific halibut, Pacific cod, arrowtooth flounder, and Kamchatka flounder that feed on Atka mackerel, other marine fishes also consume Atka mackerel. Several authors have reported that pelagic salmonids prey on Atka mackerel. Davis^ found that coho salmon (Oncor- hynchus kisiitch) ate Atka mackerel in the North Pacific Ocean. Volkov et al. (1995) found that Atka mackerel represented 30%: and 13% of the food com- position of coho salmon in the Sea of Okhotsk and the western Kamchatka areas, respectively, (jrorbimova ( 1970) stated that 25-30 mm long larvae of Atka mack- erel were found frequently in the stomachs of salmon caught in the open sea. In this study, no larval Atka mackerel were found in the stomachs of marine fishes, only juveniles or adults. NOTE Yang: The trophic role of Pleurogrammus monopterygius 1053 Table 6 Numbe rs (thous andi of Atka mackerel pop ulatior consumed bv marme fishes in three su bareas (W, C, E ) around the Aleutian Islands and the southeastern Bering Sea (SBS) in 1991. ( — 1 indicates no data. Values in parentheses are 9c total num bers. ATF = arrowtooth flounder; PCOD = Pacific cod; PH = Pacific halibut; KF = Kamchatka flounder; W = western; C = central; | E = eastern. Species (cm) W C E SBS Total ATF <20 20-39 0 0 0 0 0 0 0 0 >40 7217 (2.5) 0 0 9419(3.3) 16,639 (5.9) Total 7217 (2.5) 0 0 9419(3.3) 16,636 (5.9) PCOD <30 — — 0 0 0 30-59 3668 (1.3) 47,081 (16.6) 0 0 50,749(17.9) >60 75,358(26.51 113,249(39.9) 0 0 188,607(66.4) Total 79,026(27.8) 160,330(56.5) 0 0 239,356 (84.3) PH <50 0 0 0 0 0 50-79 0 3390 (1.2) 0 0 3390 (1.2) >80 6856 (2.4) 16,242 (5.7) 0 0 23,098 (8.1) Total 6856 (2.41 19,632 (6.9) 0 0 26,488 (9.3) KF' <20 20-39 0 0 0 0 0 0 0 0 >40 534 (0.21 0 0 999(0.4) 1533 (0.5) Total 534 (0.2) 0 0 999(0.4) 1533 (0.5) All species 93,633(33.01 179,962(63.4) 0 10,418(3.7) 284,013 ' I used ATF stomach content data and KF biomass. Table 7 Numbers (thousand bv age (vD and size (mm) of Atka mackerel consumed by the marine fishes in the Aleutian Islands area in | 1991. ATF = arrowtooth flounder; KF = Kamchatka flounder; PH = Pacific halibut; COD = Pacific cod; w = western Aleutian; | c = central Aleutian; sbs = southea stern Bering Sea AgeO Agel Age 2 Age 3 Age 4 Age 5 Age 6 Predator <131 131-206 207-261 262-302 303-333 334-354 355-370 Total ATF>40cm-w 0 0 4811 0 2406 0 0 7217 ATF>40cm-sbs 0 0 9419 0 0 0 0 9419 KF>40cm-sbs 0 0 999 0 0 0 0 999 KF>40cm-w 0 0 356 0 178 0 0 534 PH>80cm-c 0 0 8121 8121 0 0 0 16,242 PH5079cm-c 0 1130 2260 0 0 0 0 3390 PH>80cm-w 0 0 3428 1714 0 0 1714 6856 COD30-59cni-c 0 47,081 0 0 0 0 0 47,081 COD30-59cm-w 0 0 3668 0 0 0 0 3668 COD>60cm-w 0 7536 22.607 37,679 7536 0 0 75.358 COD>60cm-c 11,325 28,312 50,962 22,650 0 0 0 113,249 Total 11.325 84,059 106.631 70,-164 10.120 0 1714 284.013 Atka mackerel as predator Atka mackerel have been documented as predators of both planktonic invertebrates and teleosts in both the western and central North Pacific Ocean. Sim- enstad et al. (1977) found that planktonic crusta- ceans, hyperiid amphipods, calanoid copepods, and oikopleura (larvaceans) occurred frequently in Atka 1054 Fishery Bulletin 97(4), 1999 X 56°00-N- S4''0CCN ■ PERCENT BY WEIGHT + < 1 • 1 -25 • 26-50 • 51 -75 • 76-100 Pribilof Is. Attu I. Agattu I. Petrel Bank Buldir I. + Kiska I. -Hf- Unimak i.J r° ir 5>. Aku«nl^^. J^v^Unalaska 1. Chuginadakl^^j /(inak 1 . Q Yunaska 1 .B>^ Atkal ^ iMsna 1. ^l^ j4.„_-flhSeguam I. \ Amchllka I. ^^If Amiia I. Adakl. Figure 3 Geographic distribution of eggs cannibalized by Atka mackerel in the Aleutian Islands area in summer 1991. mackerel stomachs collected from the Amchitka Is- land area. By contrast, in this study, euphausiids were the most important food (281 by weight). Onishchik (1997) found that myctophids (57% by weight) were the most important food of Atka mack- erel in the Kuril Range area even though calanoid copepods and euphausiids occurred more frequently in stomachs. Orlov's study (1997) also showed that Atka mackerel is mainly a planktivore. He found that in the Kuril Islands area, copepods were present in 51% of Atka mackerel stomachs. Several studies (Takemura and Yamane, 1953; Zolotov and Medveditsyna, 1979; Zolotov and Tokranov, 1991) have shown that Atka mackerel eat their own eggs. I also found evidence of egg canni- balism at four locations near Kiska Island (Fig. 3). I compared the frequency of occurrence of Atka mack- erel eggs found at these locations. In three out of four hauls, females cannibalized more eggs than did males. In four hauls, males (71=7-9) exhibited 13- 439^ frequency of occurrence, and females (/!=6-8) exhibited 17-571 frequency of occurrence of egg can- nibalism. This finding suggests that male egg-guard- ing behavior, a characteristic of the species (Zolotov and Tokranov, 1991), may have inhibited the males from feeding on their own eggs. Combining stomach- content data from these four hauls (Table 8) showed that males consumed more euphausiids (271) and less calanoids (18%), whereas females consumed more calanoids ( 35% ) and less euphausiids ( 6% ) when egg cannibalism occurred. I analyzed the rest of the data (where no egg cannibalism occurred) and found that males and females fed on very similar percent- age by weights of calanoids (mainly {Neocalanus plumchrus)), euphausiids (mainly Thysanoessa spp), and larvaceans (Table 9). On the basis of these data, I hypothesize that egg-guarding behavior may cause some variations in diet of male and female Atka mackerel. All cannibalism occurred in September at depths between 80 and 170 m, indicating that the Kiska Island area may be a spawning ground for Atka mackerel in September. Zolotov and Tokranov ( 1991 ) found that, during the spawning season (from Au- gust to September), Atka mackerel eggs were the main food of rock mackerel iHcxagrammos lagocephalus), Atka mackerel, yellow Irish lord (Hemilepidotiis jordoni), and the common Irish lord (H. gilbert i). NOTE Yang: The trophic role of Pleurogrammus monopterygius 1055 Table 8 Mean percent total weight C^rW) and mean percent of requency of occurrence 9fF0) of the main prey groups of Atka mackerel (Pleurogrammus monopterygius) where egg cannibalism occurred in the Aleuti an Islands area in 1991. Prey name Male Female Total %W %FO %W %FO %W %FO Gastropoda ( snail 1 2.66 42.46 1.50 40.77 1.59 41.67 Teuthoidea (squid) 2.05 34.13 1.52 38.69 1.11 36.67 Calanoida (copepod) 18.54 66.02 34.89 77.83 21.30 71.67 Candacia columbiae (copepod) 0.98 49.06 0.80 52.23 0.74 51.67 Gammaridea (amphipod) 4.40 42.11 2.97 37.05 3.79 40.00 Themisto sp. (amphipod) 2.34 63.34 2.93 74.40 1.89 68.33 Euphausiacea (euphausiid) 27.23 49.16 6.14 42.26 19.42 46.47 Caridea (shrimp) 0.05 2.78 0.27 14.43 0.12 8.33 Majidae (spider crab) 0.48 8.68 0 0 0.20 5.00 Chaetognatha (arrow worm) 0.86 12.95 0.05 3.57 0.25 8.33 Larvacea Copelata 11.07 89.73 9.25 89.14 8.76 90.00 Gnathostomata 0.01 3.57 0 0 0.01 1.67 Zoarcidae (eelpout) 5.24 2.78 0 0 3.51 1.67 Pleurogramm us monopterygius (Atka mackerel eggs) 16.66 28.77 32.79 41.67 28.96 35.00 Fishery discards 3.92 3.57 0 0 3.05 1.67 Total prey weight 52 g 25 g 78 g Total nonempty stomachs 32 28 60 Total empty stomachs 0 0 0 Total hauls 4 4 4 Table 9 Mean percent total weight C'fW) and mean percent of frequency of occurrence ("7? FO) of the main prey groups of the Atka mackerel {Pleurogrammus monoptgerygius} that had no egg canniba ism in the Aleutian Islands area in 1991. Prey name Male Female Total %W %F0 %W %F0 %W fcFO Gastropoda (snail) 0.82 46.77 0.74 30.66 0.71 33.26 Teuthoidea (squid) 0.87 21.75 6.14 30.69 5.51 25.05 Calanoida (copepodi 30.02 60.81 22.08 60.20 23.01 58.63 Candacia columbiae (copepod) 0.32 41.52 0.53 48.36 0.46 43.43 Gammaridea (amphipod) 0.28 15.24 0.08 8.78 0.15 10.80 Themisto sp. (amphipod) 1.95 67.69 3.16 59.03 2.54 57.00 Euphausiacea (euphausiid) 26.78 63.40 31.81 68.63 30.86 63.24 Caridea (shrimp) 0.15 21.95 0.09 15.89 0.10 17.12 Majidae (spider crab) 0.03 4.04 0.05 4.37 0.03 3.53 Chaetognatha (arrow worm) 0.74 29.45 1.28 30.23 1.03 27.33 Larvacea Copelata 14.81 80.65 7.42 66.15 9.12 68.11 Gnathostomata 0.01 0.89 0.03 2.42 0.01 1.66 Myctophidae (lanternfish) 1.18 1.79 0.43 0.84 0.92 1.18 Theragra chalcogramma 2.79 2.81 4.12 2.86 2.96 2.55 Total prey weight 288 g 335 g 623 g Total nonempty stomachs 82 89 171 Total empty stomachs 5 2 7 Total hauls 14 17 17 1056 Fishery Bulletin 97(4), 1999 Atka mackerel as prey of marine mammals Many marine mammals feed on Atka mackerel. Merrick (1995) found that Atka mackerel was the most common prey identified in Steller sea lion scats from the Aleutian Islands from 1990 to 1993. Nemoto (1957) found that Atka mackerel was the preferred food of humpback whales (Megapfera novaeangliae) in the waters west of the Attn Islands and the wa- ters south of Amchitka Island. Kasamatsu and Tanaka (1992) stated that in the southwestern Hokkaido region, Pleurogfammus azonus, a conge- ner distributed mainly in the northern part of the Sea of Japan, represented SO-lOO'^ of the diet of minke whale (Balaenoptera acutorostrata). Kenyon (1965) fovmd that 85% (by volume) of the food of har- bor seals iPhoca vitiilina) in the Amchitka Islands area was Atka mackerel. Atka mackerel have also been found in the stomachs of sperm whales iPhyseter macroceph- alus) (Kawakami, 1980), killer whales iOrciniis orca) (Nishiwaki and Handa, 1958), fin whales (Balaenoplera physalus) (Nemoto, 1957), Ball's porpoise (Phocoenoides dalli) (Crawford, 1981; Perez and McAlister, 1993), minke whale (Perez and McAlister, 1993), harbor seal (Perez, 1990), northern fur seals (Callorinus ursinus) (Perez, 1990), and sea otters iEnhydra lutris) (Kenyon, 1969). Consumption of Atka mackerel at the popula- tion level by marine mammals was not estimated in this study because of insufficient data on their bio- mass, daily ration, and diet of marine mammals. Atka mackerel as prey of seabirds Seabirds are important predators in the Aleutian Islands marine ecosystem. Some seabirds feed on fish larvae or young juveniles, and some feed on zoop- lankton. Ogi (1980) analyzed the prey of 320 thick- billed murres iUria lomvia) that drowned in Pacific salmon gill nets in the oceanic waters ranging from the Kuril Islands to a region east and south of the Aleutian Islands (160'W) and found that juvenile Atka mackerel and several species of lantern fish (Myctophidae) were important prey (17% of the to- tal weight). Wehle (1983) found that Atka mackerel represented 42.3%, by occurrence, of the food of horned puffin iFratercula corniculata), and it oc- curred in 6.3% of the stomachs of tufted puffin (F. cirrhata) at Buldir Island. Hatch and Sanger ( 1992) found only two Atka mackerel ( 138 and 139 mm ) con- sumed by the tufted puffin, but Byrd et al.'^ found •I Byrd, G. V.. J. C. Williams, and R. Walder. 1992. Status and biology of the tufted puffin in the Aleutian Islands, Alaska, after a ban on salmon driftnets. U.S. Fish and Wildlife Ser\'ice. Alaska Maritime National Wildlife Refuge. Aleutian Islands Unit, PSC 486, Box .5251. FPO AP 96506-5251, Adak. Alaska. that Atka mackerel were important food of tufted puffin in 1990 at Buldir Island and Aiktak Island. Consumption of Atka mackerel at the population level by seabirds (like marine mammals) was not estimated in this study because of the lack of quan- titative data on the biomass, daily ration, and diet of many seabirds in this area. Summary This study assessed the importance of Atka mack- erel in the Aleutian Islands marine ecosystem. As a predator (mainly adults in this study), Atka mack- erel were basically zooplanktivores. Calanoid copep- ods, euphausiids, planktonic tunicates, amphipods, and other crustaceans were principal food items. Atka mackerel also consumed some benthic fishes (cottids and young-of-the-year walleye pollock) and mesope- lagic fishes (myctophids and bathylagids) and can- nibalized its own eggs. As prey, about 48,500 t of Atka mackerel (mainly age 2+) were consumed by marine fishes. Table 7 indicates that about 26% of the consumed Atka mackerel were age-3 and older. Because of insuf- ficient data, consumption of Atka mackerel at the popu- lation level by marine mammals and seabirds could not be estimated. Human beings also consumed a large amount of Atka mackerel. The total commercial catch of Atka mackerel was 26,740 t in 1991 (about 2.2% of the exploitable biomass [age 3-i-] in that year). This study provides information about predator- prey relationships between Atka mackerel and their predators (marine fishes, marine mammals, and sea- birds), and between Atka mackerel and their prey (zooplankton and other invertebrates) in the Aleu- tian Islands marine ecosystem. More information about the early life history (eggs, larvae, and juve- niles) of Atka mackerel is needed to improve our understanding of their trophic role in the marine ecosystem. Acknowledgments I would like to thank Troy Buckley, Patricia Livingston, Susanne McDermott, and Thomas Wilderbuer for their reviews of this manuscript and many sugges- tions. Robin Harrison and Mike Martin (RACE) pro- vided the survey biomass estimates and their help is appreciated. Thanks is also given to Doug Smith (REFM) for assisting with the computer programs and to Debbie Blood and Morgan Busby (RACE) for helping identify some juvenile fish, larvae, and eggs. I also want to thank three anonymous reviewers for their comments and suggestions. NOTE Yang: The trophic role of Pleurogrammus monopteiygius 1057 Literature cited Andriyashev, A. P. 1937. A contribution to the knowledge of the fishes from the Bering and Chukchi Seas. Issled. Morei 25: (Issled. Dal'nevostoch. Morei 5 1, p. 292-355. Leningrad, p. 292- 355. ITransl. by L. Lanz with N. J. Wilimovsky, 1955. U.S. Fish Wildl. Serv., Spec. Sci. Rep. 145. 81 p.] Brett, J. R., and T. D. D. Groves. 1979. Physiological energetics. //! W.S. Hoar, D.. J. Randall, and J. R. Brett (eds.). Fish physiology, vol. 8: bioenergetics and growth, p. 279-352. Academic Press, New York, NY. Crawford, T. W. 1981. Vertebrate prey o{ Phocoenoides dalli. (Dall's por- poise ), associated with the Japanese high seas salmon fish- ery in the North Pacific Ocean. M. S. thesis, Univ. Wash- ington, Seattle, WA, 72 p. Gorbunova, N. N. 1970. Spawning and development of greenlings (Family Hexagrammidae). In T. S. Rass (ed.), Greenlings, tax- onomy, biology, interoceanic transplantation. Academy of Science of the U.S.S.R. Trans. Inst. Oceanology 59:12-149. Harrison, R. C. 1993. Data report: 1991 bottom trawl survey of the Aleu- tian Islands area. U.S. Dep. Commer., NOAATech. Memo. NMFS-AFSC-12, 144 p. Hatch, S. A., and G. A. Sanger. 1992. Puffins as samplers of juvenile pollock and other for- age fish in the Gulf of Alaska. Mar. Ecol. Prog. Ser. 80: 1-14. Kasamatsu, F., and S. Tanaka. 1992. Annual changes in prey species of minke whales taken off Japan 1948-87. Nippon Suisan Gakkaishi 58(4): 637-651. Kawakami, T. 1980. A review of sperm whale food. Sci. Rep. Whales Res. Inst. Tokyo 32:199-218. Kenyon, K. W. 1965. Food of harbor seals at Amchitka Island, Alaska. J. Mammal. 46:103-104. 1969. The sea otter in the eastern Pacific Ocean. N. Am. Fauna 68, 352 p. Merrick, R. 1995. The relationship of the foraging ecology of Steller sea lions tEiimetoptas juhatus) to their population decline in Alaska. Ph, D. diss., Univ Washington, Seattle, WA. 171 p. Merrick, R. L. , M. K. Chumbley, and G. V. Byrd. 1997. Diet diversity of Steller sea lions lEumetopias jubatus) and their population decline in Alaska: a poten- tial relationship. Can. J. Fish. Aquat. Sci. 54:1.342-1348. Mito, K. 1974. Food relationships among benthic fish populations in the Bering Sea. M.S. thesis, Hokkaido Univ., Hokkaido, Japan. 135 p. Nemoto, T. 1957. Foods of baleen whales in the northern Pacific. Sci. Rep. Whales Res. Inst. Tokyo 12:33-90. Nishiwaki, M., and C. Handa. 1958. Killer whales caught in the coastal waters off Japan for recent 10 years. Sci. Rep. Whales Res. Inst. Tokyo 13:85-96 Ogi,H. 1980. The pelagic feeding ecology of thick-billed murres in the North Pacific, March-June. Bull. Fac. Fish. Hokkaido Univ 31:50-72. Onishchik, N. A. 1997. On the feeding of Atka mackerel Pleurogrammus monopterygius (Hexagrammidae) in the area of the Vityaz Ridge. J. Ichthyol. 37(8):61 1-616. Orlov, A. M. 1997. On the feeding of Atka mackerel Pleurogrammus monopterygius in the Pacific waters of the northern Kuril Islands. J. Ichthyol, 37(3):226-231. Perez, M. A. 1990. Review of marine mammal population and prey in- formation for Bering Sea ecosystem studies. U.S. Dep. Commer., NOAA Tech. Memo. NMFS F.NWC-186, 81 p. Perez, M. A., and W. B. McAlister. 1993. Estimates of food consumption by marine mammals in the eastern Bering Sea. U.S. Dep. Commer., NOAA Tech. Memo. NMFS-AFSC-14, 36 p. Simenstad, C. A., J. S. Isaksonk, and R. E. Nakatani. 1997. Marine fish communities of Amchitka Island, Alaska. In M. L. Merrit and R. G. Fuller (eds.). The envi- ronment of Amchitka Island, Alaska. U.S. Energy Research and Development Administration. TID 267-12:451-492. Takemura, Y., and T. Yamane. 1953. Notes on the food of Pleurogrammus azonus taken from the western coast of Hokkaido. Bull. Jpn. Soc. Sci, Fish. 19(2):111-117. Volkov, A. F., V. I. Chuchukalo, A. Ya. Efimkin, and I. I. Glebov. 1995. Feeding of coho salmon, Oncorhynchus kisutch. in the Sea of Okhotsk and Northwest Pacific. J. Ichthyol. 35(9): 386-391. Wehle, D. H. S. 1983. The food, feeding, and development of young tufted and horned puffins in Alaska. Condor 85:427-442. Wing, B. L. 1985. Salmon stomach contents from the Alaska troll log- book progi-am 1977-84. U.S. Dep. Commer., NOAA Tech. Memo. NMFS F/NWC-91, 43 p. Yang, MS. 1996. Diets of the important groundfishes in the Aleutian Islands in summer 1991. U.S. Dep. Commer, NOAA Tech. Memo. NMFS-AFSC-60, 105 p. Yang, M. S., and P. A. Livingston. 1986. Food habits and diet overlap of two congeneric spe- cies. Atheresthes stomias and Atheresthes evermanni , in the eastern Bering Sea. Fish. Bull. 82(31:615-623. Zolotov, O. G., and A.V. Medveditsyna. 1976. Feeding habits of the one-finned greenling in coastal waters of the north Kurile Islands. J. Ichthyol. 4(4):790- 792. Zolotov, O. G., and A. M. Tokranov. 1991. Feeding characteristics of greenlings and Irish Lords during spawning in the upper sublittoral of eastern Kamchatka. J. Ichthvol. 31(3):146-155. 1058 Ultrasonic telemetry: Its application to coral reef fisheries research Dirk C. Zeller School of Marine Biology and Aquaculture James Cook University Townsvllle, ^1 1, Australia E-mail address: DirkZelleraicueduau The importance to fisheries re- search of understanding movement patterns of fishes is increasingly being recognized (Hilborn and Walters, 1992 ). Traditionally, mark- recapture studies with external tags constituted the major method of examining movements in fishes (Shepherd, 1988). However, exter- nal tagging techniques are known to have several limitations (re- viewed by Kearney, 1989) that often cannot be addressed adequately. In particular, data obtained through conventional tagging studies are usually limited to knowledge of a single point of capture, point of re- capture, and the straight line dis- tance and time interval between these two events. Such data can be misleading because the exact dis- tance traveled and the patterns of movement are unknown. However, such patterns are of major impor- tance to current research efforts in tropical reef fisheries, including investigations of spawning aggrega- tion events (Samoilys and Squire, 1994: Zeller, 1998), and to the assess- ment of movements in relation to marine reserves (e.g. Russ and Alcala, 1996). Ultrasonic telemetry is an ideal tool to address questions of move- ment and activity patterns of fishes. It can be used effectively under circumstances that limit the use of more traditional methods, yet its suitability for fisheries re- search is only slowly being realized (Nelson, 1990). Reviews of tele- metry in the aquatic environment are provided by Harden-Jones and Arnold ( 1982), Hawkins and Urqu- hart (1983) and Nelson (1990). Ultrasonic telemetry has been little used on coral reef fishes (e.g. Holland et al., 1993; Tulevech and Recksiek, 1994); most studies have concentrated on sharks (Nelson, 1990). Recent studies, however, have successfully applied the present techniques to the serranid Plectropomus leopardus (Zeller, 1997, 1998; Zeller and Russ, 1998), the primary target species of line fisheries on the Great Barrier Reef, Australia (Kailola et al., 1993). The aims of this study were the assessment of ultrasonic telemetry in the coral reef environment and the application of the technique to P. leopardus. The first objective was the determination of a suitable transmitter placement technique and anesthetic. The second objec- tive consisted of a field evaluation of ultrasonic telemetry. The final objective comprised tracking trials of P. leopardus. Materials and methods Study location and ultrasonic telemetry equipment The study was conducted during 1993 at Lizard Island Research Sta- tion, northern Great Barrier Reef, Australia (lat. 14°40'S; long. 145° 28'E). The telemetry equipment consisted of V8-2L transmitters and a VR60 receiver linked to a direc- tional 50-80 kHz hydrophone ( Vemco Ltd, Halifax, Canada). Evaluation of fish anesthetics and transmitter placement techniques Fish (size range: 34.0 cm-52.5 cm fork length [FL]) were caught on hand lines. Aquarium facilities in- cluded 500-, 1000-, and 2000-L tanks with continuous flow-through water supply. Stocking densities did not exceed 2 fish/500 liters. Specimens were retained for an acclimation pe- riod ( 1-3 days), and were not fed for 24 hours prior to the experiments. The first concern was that of pro- curing a successful, deep anesthe- sia that allowed a gentle recovery in the fish. Three anesthetics were tested: 1) Hypnodil'^'" (Janssen Pharmaceutica, active ingredient: metomidate) was used at a concen- tration of 7 mg/L (Mattson and Riple, 1989); 2) phenoxyethanol was used at a concentration of 0.5 nil/L (Mattson and Riple, 1989); and 3) MS-222 (tricaine methanesul- fonate) was used at a concentration of 80 mg/L (Thomas and Robertson, 1991). Three methods were used to at- tach telemetry units to the animals: 1) stomach placement (force feed- ing) by means of oral insertion of the transmitter into the stomach (e.g. Holland et al., 1992)— the least intrusive method, neither resulting in external protrusion of the unit, nor requiring surgery; 2) external placement by attaching transmit- ters directly to the dorsal muscula- ture (Holland et al., 1996); and 3) internal placement by inserting transmitters into the body cavity as described by Hart and Summerfelt (1975) and Holland et al. (1993). Surgical implements were disin- fected in ethanol and soaked in Tamodine"", a fish-antiseptic solu- tion (Vetark Professional, Winches- Manu.script accepted 13 October 1998. Fish. Bull. 97(41:1058-106.5 ( 1999). NOTE Zeller: Ultrasonic telemetry: its application to coral reef fisheries research 1059 ter, UK). The incision was placed parallel to the midventral line on the left hand side of the fish, 2 cm anterior to the anus and 1-2 cm lateral of the midventral line. Three to four rows of scales were removed from the area of planned incision, and the area was cleaned with Tamodine^^'. A 2-3 cm ante- rior-posterior incision was made into the body cav- ity. Disinfected transmitters were coated in antisep- tic cream prior to careful insertion into the body cav- ity. The incision was closed with 6-8 surgical staples (Ethicon Proximate UV^') as suggested by Mortensen (1990), the area was cleaned with Tamodine^"', and the fish given an intraperitoneal injection of antibi- otic (tetracycline 50 mg/kg offish, McFarlane and Beamish, 1990), before being returned to the aquarium tanks. Aquarium experiments Owing to limited aquarium space, it was not possible to evaluate all combina- tions of anesthetics and placement methods simul- taneously. Hence two separate experiments were con- ducted. Experiment A examined the least intrusive transmitter placement technique (force feeding) with all three anesthetics. In experiment B the two intru- sive techniques (external attachment and insertion into the body cavity) were tested in conjunction with the two most suitable anesthetics. Experiment A: force feeding All three anesthetics were tested with force feeding to assess differences in retention times of transmitters due to the anes- thetic agent. It was hypothesized that use of the hyp- notic agent Hypnodil"' (metomidate) would result in the longest retention times owing to the low stress levels caused by this compound (Thomas and Robertson, 1991). Plectropomus leopardus regurgi- tate stomach contents easily, particularly when ex- posed to stressful conditions. ^ Twenty-one specimens of P. leopardus (size range: 39.0 cm-52.5 cm FL) were assigned randomly to the three anesthetic treatments in -7). Individual fish were anesthetized, tagged with T-bar anchor tags for identification, their fork length was measured, indi- vidually numbered transmitters were inserted into the stomach, and fish returned to the aquaria for recovery. Fish were fed daily. Aquaria were exam- ined for regurgitated transmitters every two hours during daytime and once during the night. The start of each inspection was taken as the maximum reten- tion period for any recovered units. ^ Davies, C. 1992. Cooperative Research Centre for Reef Re- search, James Cook University of North Queensland, Towns- ville. 4811, Austraha. Personal commun. Experiment B: external attachment and surgical insertion The suppression of a stress response due to the use of metomidate (HypnodiF^') makes this a valuable drug for the routine handling of fishes (Tho- mas and Robertson, 1991). However, slightly elevated levels of the stress hormone corticosteroid are con- sidered beneficial for resistance to trauma, thus making metomidate less suitable in situations involv- ing stress, such as surgical procedures (Thomas and Robertson, 1991). Given the nature of the intrusive manipulations performed in experiment B, the use of HypnodiF^' was discontinued. Experiment B compared two anesthetic agents (phenoxyethanol and MS-222) and two transmitter placement methods (external attachment and surgi- cal insertion) in a full factorial design. The trans- mitter placement factor consisted of five treatment levels: external and internal placement treatments, external and internal placement controls, and a rou- tine handling control. All fish received the same pre- liminary handling: fish were anesthetized, weighed (to the nearest 50 g), measured (fork length), and tagged externally. Routine-handling control fish were returned to the aquaria. The external attachment treatment consisted of attaching transmitters sensu Holland et al. (1996), and injecting an antibiotic in- traperitoneally (tetracycline 50 mg/kg offish, McFar- lane and Beamish, 1990), before fish were returned to the aquarium. The procedure lasted 5-8 minutes. External-attachment control fish underwent the same procedure, but a transmitter was not attached after puncturing the musculature, and the fish were returned to the aquaria after eight minutes. The in- ternal placement treatment consisted of surgical in- sertion of the transmitters into the body cavity as described above. The procedure took 15-20 minutes. Internal placement control fish were treated identi- cally, but no transmitter was placed in the body cav- ity prior to closure of incisions. Fish were returned to the aquaria after 20 minutes. Twenty P. leopardus were used (size range: 34.0- 50.7 cm FL), resulting in two random replicates per treatment combination. Fish were fed daily, and each individual was examined visually without handling, to record general condition, behavior, feeding and wound status. The experiment was terminated after 24 days because gas-bubble-trauma (Weitkamp and Katz, 1980) affected the behavior (lethargy and ces- sation of feeding) offish from all experimental groups. Upon termination, wounds were examined and clas- sified as aggravated (ripped, or inflamed [secondary infection] ), partially healed (40-50% wound closure), or healed (possibly minor inflammation). Subse- quently, transmitters were removed and fish were weighed. 1060 Fishery Bulletin 97(4), 1999 Data analysis Data from experiment A were ana- lyzed for differences in retention times between an- esthetic agents by using a single-factor analysis of covariance (ANCOVA). Fork length was used as a covariate to account for any effect of size of fish on the retention times. Assessment of experiment B was based on the observations made during and after the experiment. The change in weight during experiment B was analyzed by using a two-factor analysis of co- variance (ANCOVA). Occurrence of gas-bubble- trauma was treated as a covariate. Underlying as- sumptions were evaluated prior to analyses (Sokal andRohlf, 1981). Field evaluation of ultrasonic telemetry in the coral reef environment The extensive use of radio telemetry in wildlife re- search has produced procedures to assess the accu- racy (bias and precision) of telemetry reception (White and Garrott, 1990). However, no such proce- dures appear to be published for ultrasonic telemetry. Given the paucity of studies with ultrasonic telemetry in the coral reef environment, a standardized field test using stationary transmitters was undertaken. Given that wind and sea-state influence the detect- ability of sound transmission in water (Jellyman et al., 1996), any effect of direction of prevailing wind on detectability or directional bias of observed sound signals needed to be assessed. Thus, the objectives of this field evaluation were 1) to establish the accu- racy of directional bearings in relation to prevailing wind direction; 2) to evaluate observer bias and pre- cision of bearings in relation to wind direction; and 3) to determine the optimal angle for cross-bearings, and the optimal distance between tracking boat and signal in order to obtain position estimates that mini- mize error polygons. 21 positions for the three transects. Replicated com- pass bearings to the transmitter were obtained for each position by using the following protocol: the field of view of the observer was restricted to the tracking receiver for the duration of the experiment; an as- sistant anchored the boat at the marker buoys in random order and recorded the true compass bear- ing to the transmitter. The observer determined the perceived maximum directional signal strength by using the directional hydrophone. The correspond- ing compass bearing of the directional hydrophone (observed bearing) was recorded. The direction of the hydrophone was changed haphazardly by the assis- tant and the procedure was repeated. Six replicate bearings were taken at each of the seven distances on the three transects. The procedure was repeated by the second observer for a total of 252 observed bearings. Data analysis Data were assessed for accuracy, bias, and precision (Wliite and Garrott, 1990). Bearing error polygons were determined by using the largest and smallest directional bearing observed at each distance-marker buoy for any combination of two transects (0''-45°, 45°-90°, and 0°-90°). Thus, the error polygon parameters obtained (polygon area and maximum diagonal dimension) represented the larg- est possible error polygon estimates. Only bearing pairs from equidistant locations were used for poly- gon determination. Statistical analyses included Kruskal-Wallis nonparametric ANOVA and paired t- test. All data were examined for violations of under- lying assumptions prior to analysis (Sokal and Rohlf, 1981) and data log^Q transformed where applicable. Given that the angular scale of measurement repre- sented only part of a circle, and absolute direction of angles was of no interest, data were treated as lin- ear (Cain, 1989). Experimental methods An ultrasonic transmitter (V8-2L, Vemco Ltd) was attached 15 cm above the substratum to a surface buoy in 4-6 m water depth. Thi-ee 200-m transects were run fi-om the moored trans- mitter in relation to the prevailing wind direction: 0": transmitter located directly upwind from any position on the transect. 45": transmitter located at 45° downwind from the prevailing wind direction. 90°: transmitter located at right angle to the prevail- ing wind direction. Marker buoys were positioned at 25-m intervals along each transect, starting at 50 m from the trans- mitter location (i.e. from 50 to 200 m), resulting in Field tracking trials To verify the findings from the aquarium study, the two most suitable long-term transmitter placement techniques (external and internal placement) were tested under field conditions (MS-222 used as anes- thetic). Two P. leopardus (43.1 cm and 58.9 cm FL) were equipped with external transmitters and re- leased at their capture site after a 15-min recovery period. In the second field trial, internal transmit- ters were placed in two specimens (59.0 cm and 42.9 cm FL). Fish were released at the capture site after recovery from anesthesia. The basic tracking technique described by Holland et al. (1985, 1992) was used. Exact positions offish equipped with transmitters were determined by visual NOTE Zeller: Ultrasonic telemetry: its application to coral reef fisheries research 1061 triangulation (White and Garret, 1990) by using identifiable landmarks and reef fea- tures. Triangulation involves the estimation of the location of a transmitter by using two or more directional bearings obtained from a known location (White and Garrot, 1990). Visual triangulation uses visible land and reef features that can be identified from maps or aerial photographs, in conjunction with directional bearings to identify the lo- cation of the transmitter Results Evaluation of fish anesthetics and transmitter placement techniques Experiment A: force feeding Gastric re- tention times of force-fed transmitters ranged from 18 to 216 hours; longer re- tention periods were documented for larger fish (Fig. 1, r=0.621, n=21). Mean retention times differed between anes- thetic agents used (ANCOVA, F., ^-=4.762, P=0.023). The use of MS-222 resulted in a shorter retention time compared with ei- ther phenoxyethanol (SNK, P=0.004) or metomidate (SNK, P=0.001). Mean reten- tion times did not differ with the use of either phenoxyethanol or metomidate (SNK, P=0.216). On average, fish anesthe- tized with MS-222 retained the transmit- ters for 42.0 hours (±9.7 SE), in contrast with fish anesthetized with phenoxy- ethanol (118.0 hours [±17.2 SE]) and metomidate (147.4 hours [±23.8 SE]). Experiment B: external attachment and surgical insertion The change in weight offish over the duration of the experiment differed significantly between transmitter placement treatments (ANCOVA, F^g= 5.309, P=0.018). Fish carrying external transmitters displayed a substantial re- duction in weight compared with fish in all control groups that gained weight, whereas the internal treatment group showed no change in mean weight during the experiment (Fig. 2). Fish with exter- nal transmitters were observed spending a lot of time rubbing the attached trans- mitter against the substratum. Further- more, most fish started feeding within two days cept for fish in the external treatment group 225 D 200 0 G D 175 □ g 150 ■1 125 c o c 100 0) ■5 °: 75 o o □ o 0 A D ° D o D Metomidate 50 A A MS-222 25 A A A o Phenoxyethanol A A 0 < 35.0 37.5 40.0 42.5 45.0 47.5 50.0 52.5 55.0 Fork length (cm) Figure 1 Correlation of gastric retention time of force-fed transmitters versus fork length of P. leopardus, as determined in experiment A (n=21, ;=0.62). The three anesthetic agents are indicated. Lij c (0 O 250 200 150 [ 100 ■ 50 '- 0 -50 ■100 -150 •200 - -250 -300 EC ET Treatment IC IT Figure 2 Change in mean weight (before treatment to after treatment) of P. leopardus as determined in experiment B. Note: negative weight indi- cates a loss in weight and vice versa. Presented are means (g ±SEl. Treatments:' C = routine handling control; EC = external attachment control; ET = external attachment treatment; IC = internal placement control; IT = internal placement treatment, n = 20. , ex- those anesthetized with phenoxyethanol in the in- and ternal control group ( Table 1 ). 1062 Fishery Bulletin 97(4), 1999 Table 1 Observations for aquarium experiment B. Treatment: C = routine handling control; EC = external attachment control; ET = external attachment treatment; IC = internal placement control; IT = internal placement treatment. Anesthetic; MS = MS-222; P = phenoxyethano . " — " indicates that specimen never res umed feeding after treatment Type of Level of anesthesia Day of Affected by Treatment anesthetic achieved first feeding gas -bubble-trauma Wound status C MS deep 2 Y — C MS deep 2 N — c P twitching 2 N — c P deep 2 N — EC MS deep — Y partially healed EC MS deep 2 N healed EC P deep — Y partially healed EC P twitching 2 Y healed ET MS deep 10 Y aggravated ET MS deep — Y aggravated ET P deep 10 N aggravated ET P twitching 7 N aggravated IC MS deep 2 N healed IC MS deep 2 N healed IC P deep 7 Y healed IC P deep 7 N healed IT MS deep 2 N healed IT MS deep 2 N healed IT P deep 2 N healed IT P twitching — Y healed Observational records illustrated some differences in the effects of the anesthetics. All fish anesthetized with MS-222 attained deep anesthesia, whereas 407f offish exposed to phenoxyethanol did not (Table 1 ). Tliis resulted in twitching of the animal during treatment. Occasional cramp-like convulsions were also ob- served in some of these specimens during the early recovery period. Examination of placement wounds after the ter- mination of the experiment revealed consistent pat- terns (Table 1). The incisions made into the body cavi- ties of internal control and internal treatment fish were healed (some minor inflammations did exist). The punc- ture wounds through the dorsal musculature of fish with externally attached transmitters were aggravated and in some cases enlarged due to repeated attempts to dislodge the transmitters. Wounds on external con- trol fish were either healed or partially healed, and there were no further signs of aggravation. Field evaluation of ultrasonic telemetry in the coral reef environment Evaluation of bearing errors (true bearing-observed bearing) by transect (0°, 45° vs. 90°) indicated a sig- nificant difference in mean bearing error between transects (Ki-uskal-Wallis ^2 252=22.102, P= 0.000). The 90° transect displayed the least bias (mean=- 1.50° ±0.7163° SE), the 0° transect had slightly larger, positive bias (mean=2.09° ±1.077°SE), and the 45° transect showed strong, negative bias (mean=- 5.43°±1.021°SE). Owing to the differences in bias and accuracy be- tween transects, observer differences in bias were examined separately for each transect. Bias differed between observers for the 0° transect (Kruskal-Wallis //j y4=6.238, P=0.013); observer two showed positive bias compared with observer one (Fig. 3). Observers did not differ significantly in bias for the 45° transect (Kruskal-Wallis H^ y,,=0.209, P=0.648, Fig. 3), or the 90° transect, (Kruskal-Wallis H, ,^=2.603, P=0.107, Fig. 3). Paired ^-tests were used to compare precision esti- mates (SD) between observers for each transect sepa- rately. No significant observer difference in precision was detected: 0° (^g=1.2599, ^=0.2545), 45° (fg= 1.5819, P=0.1648) and 90° (