U.S. Department of Commerce Volume 90J Number 1 January 1992 ine Biological Laboratory LIBRARY AY 5 mz f"W(Wds Hole, Mass. U.S. Department of Commerce Barbara Hackman Franklin Secretary National Oceanic and Atmospheric Administration John A. Knauss Under Secretary for Oceans and Atmosphere National Marine Fisheries Service William _W. Fox Jr. Assistant Administrator for Fisheries ^^ATci 0< *' The Fishery Bulletin (ISSN 0090-0656) is published quarterly by the Scientific Publications Office, National Marine Fisheries Service, NOAA, 7600 Sand Point Way NE, BIN C15700, Seattle, WA 98115-0070. Second class postage is paid in Seattle, Wash., and additional offices. POSTMASTER send address changes for subscriptions to Fishery Bulletin, Super- intendent of Documents, Attn: Chief, Mail List Branch, Mail Stop SSOM, Washington, DC 20402. Although the contents have not been copyrighted and may be reprinted entire- ly, reference to source is appreciated. The Secretary of Commerce has deter- mined that the publication of this period- ical is necessary in the transaction of the public business required by law of this Department. Use of funds for printing of this periodical has been approved by the Director of the Office of Management and Budget. For sale by the Superintendent of Documents, U.S. Government Printing Office, Washington, DC 20402. Subscrip- tion price per year: $16.00 domestic and $20.00 foreign. Cost per single issue: $9.00 domestic and $11.25 foreign. See back page for order form. D D 3QaDD(SGDDD Scientific Editor Dr. Linda L. Jones National Marine Mammal Laboratory National Marine Fisheries Service, NOAA 7600 Sand Point Way NE Seattle, Washington 981 15-0070 Editorial Committee Dr. Andrew E. Dizon National Marine Fisheries Service Dr. Charles W. Fowler National Marine Fisheries Service Dr. Richard D. Methot National Marine Fisheries Service Dr. Theodore W. Pietsch University of Washington Dr. Joseph E. Powers National Marine Fisheries Service Dr. Tim D. Smith National Marine Fisheries Service Dr. Mia J. Tegner Scripps Institution of Oceanography IVIanaging Editor Nancy Peacocl< National Marine Fisheries Service Scientific Publications Office 7600 Sand Point Way NE, BIN CI 5700 Seattle, Washington 981 15-0070 The Fishery Bulletin carries original research reports and technical notes on investiga- tions in fishery science, engineering, and economics. The Bulletin of the United States Fish Commission was begun in 1881; it became the Bulletin of the Bureau of Fisheries in 1904 and the Fishery Bulletin of the Fish and Wildlife Service in 1941. Separates were issued as documents through volume 46; the last document was No. 1 103. Begin- ning with volume 47 in 1931 and continuing through volume 62 in 1963, each separate appeared as a numbered bulletin. A new system began in 1963 with volume 63 in which papers are bound together in a single issue of the bulletin. Beginning with volume 70, number I, January 1972, the Fishery Bulletin became a periodical, issued quarterly. In this form, it is available by subscription from the Superintendent of Documents, U.S. Government Printing Office, Washington, DC 20402. It is also available free in limited numbers to libraries, research institutions, State and Federal agencies, and in exchange for other scientific publications. U.S. Department of Commerce Seattle, Washington Volume 90 Number 1 January 1992 Fishery Bulletin Contents 1 Buckland, Stephen T., Karen L. Alejandro A. Anganuzzi Estimating trends in abundance of dolphins associated with tuna in the eastern tropical Pacific Ocean, using sightings data collected on commercial tuna vessels 13 Collette, Bruce B., and Gary B. Gillis Morphology, systematics, and biology of the double-lined mackerels [Grammatorcynus, Scombridae) 54 Douglas, Michael E., Gary D. Schnell, Daniel J. Hough, and William F. Perrin Geographic variation in cranial morphology of spinner dolphins Stenella longirostns in the eastern tropical Pacific Ocean 77 Gall, Graham A.E., Devin Bartley, Boyd Bentley, Jon Brodziak, Richard Gomulkiewicz, and Marc Mangel Geographic variation in population genetic structure of Chinook salmon from California and Oregon 101 Hunter, J. Roe, Beverly J. Macewicz, N. Chyan-huei Lo, and Carol A. KImbrell Fecundity, spawning, and maturity of female dover sole Microstomus paaficus, with an evaluation of assumptions and precision 1 29 Kendall, Arthur W. Jr., and Toshikuni NakatanI Comparisons of early-life-history characteristics of walleye pollock Theragra chalcogramma in Shelikof Strait, Gulf of Alaska, and Funka Bay, Hokkaido, Japan 139 McShane, Paul E. Exploitation models and catch statistics on the Victorian fishery for abalone Haliotis rubra Fishery Bulletin 90(1). 1992 147 Sissenwine, Michael P., and Pamela M. Mace ITQs in New Zealand: The era of fixed quota in perpetuity 161 Stoner, Allan W., Veronique J. Sandt, and Isabelle F. Boidron-Metairon Seasonality in reproductive activity and larval abundance of queen conch Strombus gigas 171 Wainwright, Thomas C, David A. Armstrong, Paul A. Dinnel, Jose M. Orensanz, and Katherine A. McGraw Predicting effects of dredging on a crab population: An equivalent adult loss approach Notes 183 Chen, Weihzong, John J. Govoni, and Stanley M. Warlen Comparison of feeding and growth of larval round herring Etrumeus teres and gulf menhaden Brevoortia patronus 190 D'Amours, Denis, and Francois Gregoire Analytical correction for oversampled Atlantic mackerel Scomber scombrus eggs collected with oblique plankton tows 1 97 Rajaguru, Arjuna, and Gopaisamy Shantha Association between the sessile barnacle Xenobalanus globiapitis (Coronulidae) and the bottlenose dolphin Tursiops truncatus (Delphinidae) from the Bay of Bengal, India, with a summary of previous records from cetaceans 203 Safford, Susan E., and Henry Booke Lack of biochemical genetic and morphometric evidence for discrete stocks of Northwest Atlantic herring Clupea harengus harengus 211 Stergiou, Konstantinos I. Variability of monthly catches of anchovy Engraulis encrasicolus in the Aegean Sea Abstract.- We summarize the methods for estimating relative abun- dance of seven dolphin stocks in the eastern tropical Pacific Ocean using sightings data collected on commer- cial tuna vessels by trained observ- ers, developed by Buckland and Anganuzzi (1988a) and Anganuzzi and Buckland (1989). Their estimates of relative abundance, which may show large year-to-year fluctuations, are smoothed to provide estimates of the underlying trend in dolphin abun- dance between 1976 and 1988. The bootstrap method provides estima- tion of precision in a way that allows trend estimates to be used for man- agement purposes, without the need to assume that trends in abundance are linear. Concerns about the valid- ity of the estimates are addressed. Estimating trends in abundance of dolphins associated witli tuna in the eastern tropical Pacific Ocean, using sightings data collected on commercial tuna vessels Stephen T. Buckland Karen L. Cattanach SASS Environmental Modelling Unit, MLURI Craigiebuckler, Aberdeen AB9 2QJ. United Kingdom Alejandro A. Anganuzzi Inter-American Tropical Tuna Commission 8604 La Jolla Shores Drive, La Jolla, California 92093 Manuscript accepted 27 November 1991. Fishery Bulletin, U.S. 90:1-12 (1992). Incidental mortality of dolphins in the tuna fishery in the eastern trop- ical Pacific since 1959 has been suf- ficient to affect abundance of stocks of at least two species of dolphin: the spotted dolphin Stenella attenuata and the spinner dolphin S. longiros- tris (Smith 1983). Although there is less information available on stocks of the common dolphin Delphinus delphis, mortality estimates (e.g., Hall and Boyer 1988) suggest that abundance of stocks of this species may also have been reduced. To mon- itor possible effects of incidental mor- tality on the size of dolphin stocks, several attempts to estimate abun- dance have been made, usually apply- ing line-transect methodology to data collected on either commercial tuna vessels ("tuna vessel data") or re- search vessels ("research vessel data") or both. Holt and Powers (1982) and Holt (1985, 1987) consid- ered analyses of research vessel data alone, and of tuna vessel data com- bined with research vessel data. More recently. Holt and Sexton (1989, 1990a, b) analyzed data from research vessels alone. Tuna vessel data alone were analyzed by Ham- mond and Laake (1983), by Polacheck (1987), by Buckland and Anganuzzi (1988a), and by Anganuzzi and Buck- land (1989). The tuna vessel data are collected by scientific technicians placed by two organizations onboard commer- cial tuna purse seiners. The Inter- American Tropical Tuna Commission (lATTC) places technicians on vessels of the international fleet (including U.S. -registered vessels), and the National Marine Fisheries Service (NMFS) of the United States places technicians on U.S. -registered ves- sels only. Data were first collected by NMFS in 1974, and by lATTC in 1979. Tuna vessel data provide a large database, with regular coverage of a substantial portion of the area oc- cupied by the dolphin stocks. How- ever, due to the nature of the fishery operations, the assumptions neces- sary for line-transect sampling to yield unbiased estimates of absolute abundance are often violated. There- fore, analytic procedures should as far as possible be insensitive to those violations. We summarize here the procedures of Buckland and Anga- nuzzi (1988a), as modified by Anga- nuzzi and Buckland (1989). Since these procedures are unlikely to re- move all biases, the estimates should Fishery Bulletin 90(1). 1992 be treated as indices of relative abundance, rather than estimates of absolute abundance of the stocks. The definition of a stock, and its boundaries, is problematic, but we follow the recommendations of Au et al. (1979), for reasons stated by Anganuzzi and Buckland (1989), except in two cases. A more southerly southern bound- ary was found to be necessary for the southern offshore stock of spotted dolphins (Anganuzzi et al. 1991), and we adopt the recommendation of Perrin et al. (1991) to combine the northern and southern whitebelly stocks of spinner dolphins. We also derive estimates for pooled offshore stocks of spotted dolphins and pooled stocks of common dolphins, since they are not differentiable in the field. Buckland and Anganuzzi (1988a) provided three types of test for assessing whether abundance of a stock had changed over time. For several stocks, the tests failed to provide a clear indication of recent changes, since the occasional large fluctuation in an- nual estimates indicated that there were significant changes in abundance that were biologically implaus- ible. We present here a method of smoothing the sequence of estimates of relative abundance. Used in conjunction with the bootstrap, it yields a simple method of assessing change over time which does not require that trends are assumed to be linear, and which does not yield biologically implausible rates of change. Edwards and Kleiber (1989) have questioned the validity of estimating trends in abundance from sight- ings data collected on commercial tuna vessels. We carry out a simple simulation study to assess their assertions, and compare the relative abundance esti- mates calculated from tuna vessel data with those calculated from research vessel data for the years 1986-89, for which data from both sources are available. Methods The number of dolphins A^ in an area for a given stock and year is estimated by N = A ■ S ■ D where A is the size of the area, s is the estimated average school size for the stock in area A, and D is the estimated density of schools in area A. The line-transect method provides the estimate D (Burnham et al. 1980). Suppose schools farther than a distance w from the trackline are discarded from the analyses. Then D = 2L (1) where n is the number of schools detected in the area that are within the truncation distance w, /(O) is the estimated probability density function of the n perpendicular distances, evaluated at perpendicular distance zero, and L is the total length of transect in nautical miles within the area. If we define the encounter rate E to be the expected number of sightings detected within m' of the trackline per nautical mile of search, then its estimate is given by E = nIL. Hence, and D N E-m Ef{0)-s-A (2) (3) UD and N were estimates of absolute abundance, then the following assumptions would be required: (i) Within each area or stratum, either the search effort of the tuna vessels is random or the dolphin schools are randomly distributed; (ii) any movement of schools is slow relative to the speed of the vessel, at least before detection; (iii) all schools on or close to the trackline are detected and identified; (iv) sighting distances and angles are measured with- out error; (v) sightings of schools are independent events; (vi) school size is recorded without error, and for mixed schools percent of each species is recorded without error; (vii) probability of detection of a school is independent of its size, at least out to perpendicular distance w. If the estimates are used solely as indices of relative abundance, as here, then any or all of the above assumptions may fail without invalidating the esti- mates, provided that bias arising from the failure of an assumption is consistent across time. Even this pro- viso may be relaxed when trends in abundance over a long sequence of years are estimated; in this case it is merely necessary to assume that bias shows no trend with time. Catch-per-unit-effort methods for estimating relative abundance are known to show trends in bias over time in some instances, due to increased efficiency of vessels (Cooke 1985). We attempt to avoid such prob- lems by incorporating a parameter that measures the Buckland et al.: Estimating abundance of tuna-associated dolphin stocks in the eastern tropical Pacific efficiency of search of the tuna vessels. This parameter, the effective search width, is estimated using line- transect theory. It may be interpreted as twice the distance at which the number of undetected dolphin schools closer to the vessel is equal to the number of detected schools further from the vessel, and is there- fore the effective width of the strip of ocean searched by the vessel. As efficiency of the fleet to detect dolphin schools increases (e.g., through the use of helicopters, high-resolution radar, etc.), the effective search width increases, and bias in abundance estimates should re- main unaffected. We adopt a strategy of reducing bias as much as possible, so that the effect of any trend in bias over time on estimated trends in abundance is minimized. To estimate the different components of the estimator of Equation (3), separate stratification schemes are ap- plied for encounter rate, effective search width, and school size. In stratifying for a given component, our aim is to define strata such that each stratum is relatively homogeneous with respect to that compo- nent, so that non-random search effort and non-random distribution of schools generate only small bias in any given stratum. Crude encounter rates, average school sizes, and average detection distances are estimated by 1° square. Where data are insufficient, the crude estimates are smoothed, and the same smoothing pro- cedure interpolates for squares in which there was no tuna vessel effort. These estimates are used to allocate 1 ° squares to strata, yielding the separate stratifica- tions for encounter rate, school size, and effective search width, respectively. Full details are given by Anganuzzi and Buckland (1989). Thus the problem of abundance estimation has been reduced to three simpler problems: For a random point in the stock area, the expectations of encounter rate, school size, and effective search width are estimated, and the three estimates are multiplied together to ob- tain the final abundance estimate. Lack of indepen- dence between the three estimates does not bias the overall estimate, and independence is not assumed when estimating variance. A nonparametric bootstrap technique is used to obtain variances. The resampling unit in the bootstrap is the individual cruise, and for each bootstrap replicate the full estimation procedure is applied, thus generating bootstrap estimates of abun- dance. The sample variance of these estimates yields the required variance estimates, and confidence inter- vals are obtained by the percentile method. (See Buck- land and Anganuzzi 1988a, for details.) Bias arising from rounding errors in the recorded sighting distances r and angles 9 is reduced by smear- ing the data, using the method favored by Buckland and Anganuzzi (1988b). The recorded location of each school relative to the tuna vessel at the time of detection is defined by r and 9, and that location is "smeared" over the sector defined by r • (1 ± d ) and 9 ± ^12. to allow for inaccuracy in the recorded values. The smearing parameters d and I are estimated from the data. When a small sighting angle is rounded to zero, the calculated perpendicular distance is zero, giving a spurious spike in the perpendicular distance distribution at zero distance. Smearing yields more robust estimation by removing or reducing this spike. Here we take the estimates of Anganuzzi and Buck- land (1989) and of Anganuzzi et al. (1991) and attempt to estimate the underlying trends in dolphin abundance by smoothing them. Various smoothing methods such as moving averages, running medians, and polynomial regression were investigated (Smith 1988). The chosen method was a compound running median known as "4253H, twice" (Velleman and Hoaglin 1981), which is constructed as follows. Suppose that {X{t )}, ^ = 1, . . . , A'^, is a time-series of length A^, and let {5,(0} be a smoothed version of it, found by calculating an i -period running median. We can construct compound smoothing methods such as {Sijit)}, which is simply {Sj{Si{t))}. Thus, a 4253 run- ning median method smooths a time-series using a 4-period running median, which is in turn smoothed by a 2-period running median, smoothed again by a 5-period running median, and then by a 3-period running median (i.e., {54253(0} = {5'3(S5(S2(S4(0)))})- Near the endpoints, where there are not enough values surrounding a point to be smoothed using the spe- cified running median, a shorter-period running median may be used. The endpoints of the resultant time-series are calculated by estimating X(0) and X(N + l), the "observed" values at t = 0 and t=N + l, and then calculating 54253(1) = median {1(0), X{1), 54,53(2)} and 54253(iV) = median {S4253(A^-1), XiN), X{N + 1)}. X(0) is found by extrapolating from the straight line which passes through the smoothed values att=2 and i = 3, i.e., 1(0) = 3 -54953(2)- 2 -54953(3); similarly, X{N + 1) = 3- 54953(iV - 1) - 2 - 54253(^ - 2). The H in "4253H, twice" denotes a linear smoothing method commonly used with running medians, which is known as Banning. It is a 3-period weighted mov- ing average iort=2,...,N-l, with weights {0.25, 0.5, 0.25}. The endpoints remain unchanged. The pattern of the time-series may be recovered by calculating the residuals of the series (i.e., the differ- ences between the smoothed and unsmoothed esti- mates), smoothing the residual series using the same method as for the time-series, and then adding the smoothed values of the residuals to the smoothed Fishery Bulletin 90(1), 1992 values of the series. This is known as smoothing "twice." For example, if we define the residuals of the time-series smoothed by 4253H to be {E(t )} = {X(t ) - •54253 (i )}. then the values of the times-series smoothed by "4253H, twice" can be defined by {•S4253H, twice (0} = {54253h(0 + 'S4253h(-E'(0)}- Thus the "4253H, twice" running median method uses a 4253 running median to smooth the time-series, estimates the endpoints of the smoothed series, and then smooths the resultant series by Manning. The residuals of the series are calculated and are also smoothed, using the same method as above. The smoothed values of the residuals are then added to the smoothed values of the time-series to produce a time- series smoothed by "4253H, twice." The advantage of using running medians is that the magnitude of an extreme estimate does not affect the resultant smoothed time-series. The above method is sufficient- ly complex that its behavior cannot be readily under- stood. However, simpler methods were found to suf- fer from one or more of the following shortcomings: Estimated trends were not always smooth; implausible rates of change were sometimes indicated; trends near the start or end of the sequence of estimates were often poorly estimated. Nonparametric bootstrap replicates are generated as described by Anganuzzi and Buckland (1989). We select here the bootstrap estimates that correspond to an 85% confidence interval for relative abundance in each year. The rationale for the choice of confidence level is that if two 85% confidence intervals do not overlap, the difference between the corresponding relative abun- dance estimates is significant at roughly the 5% level (P<0.05); whereas if they do, the difference is not significant (P>0.05). If the abundance estimates are assumed to be lognormally distributed, each with the same coefficient of variation, then the exact confidence level that gives this property is 83.4%. If one estimate has twice the coefficient of variation of the other, the confidence level increases slightly to 85.6%. Thus a choice of 85% makes some allowance for variability in the coefficient of variation. . For each abundance estimate, 79 bootstrap replicates are run, so that the 6th smallest and 6th largest boot- strap estimates provide an approximate 85% confi- dence interval (Buckland 1984). If this procedure is carried out independently for each year, confidence intervals are wide. Provided the assumed stock area spans the whole range of the stock, numbers of dolphins within it are unlikely to vary greatly in successive years, and a procedure that calculates confidence in- tervals for a given year incorporating information from years immediately preceding and following that year is more informative. For a given stock, we achieve this by carrying out one bootstrap replication for each year that a relative abundance estimate is available. These estimates are smoothed using the routine described above, and the process is repeated 79 times. For each year, the 6th smallest and 6th largest smoothed estimates provide approximate 85% confidence limits. We use the sequence of medians of the smoothed boot- strap estimates (i.e., the 40th estimate of each ordered set of 79) as the "best" indicator of trend, so that it is calculated in a comparable manner to the confidence limits. Larger numbers of bootstrap replicates are preferable, but available computer power was limited. Repeat runs for the northern offshore stock of spotted dolphins were carried out, to assess the Monte Carlo variability. By using overlapping confidence intervals to test for a difference between years, independence between smoothed estimates for different years is assumed. Given the strong positive correlation in the smoothed estimates between successive years, the test is unlike- ly to detect a large change between one year and the next, but should be reliable for detecting trends over a period of perhaps five or more years, for which cor- relations between smoothed estimates are small. Results Figures 1-10 show the estimates of underlying trend for each of the main stocks associated with tuna in the eastern tropical Pacific Ocean. Since stock boundaries and stock identity are both uncertain, we also show trend estimates after pooling data from stocks that are not differentiable in the field. The broken horizontal lines in these plots correspond to the upper and lower 85% confidence limits for the 1988 relative abundance estimate. Years for which the entire confidence inter- val lies outside the region between the broken horizon- tal lines show a relative abundance significantly different from that for 1988. Because the smoothed estimate for the first or final year of a sequence can be poor, we show the unsmoothed estimate and cor- responding 85% confidence limits for the first and last year on each plot. Figures 1 and 2 show estimated trends for northern offshore spotted dolphins, with and without the abnor- mally low 1983 estimate, which corresponded with a very strong El Nino event. It is clear that the 1983 estimate affects the smoothed estimate of trend, but its effect is no greater than if it had been just smaller than the 1984 estimate. Thus abnormal estimates may be more safely retained when using this procedure, and subjective decisions of whether to treat an estimate as an outlier are avoided. Buckland et al : Estimating abundance of tuna-associated dolphin stocks in the eastern tropical Pacific 7000 -6000 1 i Jsooo Itooo 1 "§3000 a 2000 1000 Q^^^ D 1975 1976 1977 1973 1979 1980 1981 1982 1983 1984 1985 1986 1987 1988 1989 Figure 1 Smoothed abundance trends of northern offshore stock of spotted dolphin Stenella attenuata in the eastern tropical Pacific. Broken lines indicate approximate 85% confidence limits. Horizontal lines correspond to 85% confidence limits for the 1988 estimate. If lower limit lies above upper limit for an earlier year, abundance has increased significantly between that year and 1988 (P< 0.05); if upper limit lies below lower limit for an earlier year, abundance has decreased significantly. The estimated trend from Figure 1 is downwards until around 1983. Estimated abundance in 1976 and 1977 was significantly higher than in 1988 (P<0.05), but there is some evidence of a recovery between 1983 and 1988 (P<0.05). Thus northern offshore spotted dolphins appeared to decrease through the 1970s and early 1980s, with numbers remaining stable or increas- ing since. Figure 3 suggests there may have been a marked decline in numbers of southern offshore spotted dol- phins since the late 1970s. The smoothed 1988 estimate is significantly lower than the smoothed estimates for 1977 and 1978, but there is evidence of an increase since 1986 (P<0.05), after a relatively high unsmoothed estimate for 1989. As shown by Anganuzzi et al. (1991), southern offshore spotted dolphins appear to occupy appreciably different regions from one year to another, and the extent of mixing with northern offshore spotted dolphins remains unclear. We therefore believe that trend estimates for this stock are unreliable. The estimated trends obtained by pooling data from the off- shore stocks are shown in Figure 4. The estimates are dominated by the data from the larger northern off- shore stock, and the plot is similar to Figure 1. The 1988 smoothed relative-abundance estimate is signifi- cantly higher than the 1983 and 1984 estimates, and significantly lower than all estimates preceding 1979. 1975 1976 1977 1978 1979 1980 1981 1982 1983 1984 1985 1986 1987 1988 1989 Figure 2 Smoothed abundance trends of northern offshore stock of spotted dolphin Stenella attenuata in the eastern tropical Pacific, excluding 1983 estimate. Broken lines indicate approx- imate 85% confidence limits. See Figure 1 for more details. 1975 1976 1977 1978 1979 1980 1981 1982 1983 1984 1985 1986 1987 1988 1989 Figure 3 Smoothed abundance trends of southern offshore stock of spotted dolphin Stenella attenuata in the eastern tropical Pacific. Broken lines indicate approximate 85% confidence limits. See Figure 1 for more details. Figure 5 suggests that the eastern spinner dolphin might have had a pattern of change similar to the northern offshore spotted dolphin, although estimated abundance in the late 1980s is roughly equal to that in the mid-1970s, so depletion between 1975 and 1983 may have been less than for northern offshore spotted dolphins. The 1988 smoothed estimate is just signifi- cantly higher than the smoothed estimates for 1981 and 1982 (P<0.05). Fishery Bulletin 90(1). 1992 1975 1976 1977 1978 1979 1980 1981 1982 1983 1984 1985 1986 1987 1988 1989 Figure 4 Smoothed abundance trends of pooled northern and southern offshore stocks of spotted dolphin Stenella attenuata in the eastern tropical Pacific. Broken lines indicate approximate 85% confidence limits. See Figure 1 for more details. 1975 1976 1977 1978 1979 1980 1981 1982 1983 1984 1985 1986 1987 1988 1989 Figure 6 Smoothed abundance trends of whitebelly stock of spinner dolphin Stenella longirostris in the eastern tropical Pacific. Broken lines indicate approximate 85% confidence limits. See Figure 1 for more details. 1975 1976 1977 1978 1979 1980 1981 1982 1983 1984 1985 1985 1987 1998 1989 Figure 5 Smoothed abundance trends of eastern stock of spinner dolphin Stenella longirostris in the eastern tropical Pacific. Broken lines indicate approximate 85% confidence limits. See Figure 1 for more details. soo •;r I ^ .-' □ £ 500 £ X ? ' / '"' D-^ \ i 400 / / =^^-_^ X ^ / , ^^ ^ """-^^ 1 / ^/"■"^ ° ■~- * / ^ ^ i 200 \ 1975 1976 1977 1978 1979 1980 1981 1982 1983 1984 1985 1986 1987 1988 1989 Figure 7 Smoothed abundance trends of northern stock of common dolphin Delphinus delphis in the eastern tropical Pacific. Broken lines indicate approximate 85% confidence limits. See Figure 1 for more details. The estimated trend for whitebelly spinner dolphins (Fig. 6) is similar to that for eastern spinner dolphins and northern offshore spotted dolphins. There is some evidence that abundance in 1988 was higher than in 1982 (P=0.05), but no other comparisons with 1988 are significant. The 1982 smoothed estimate is significantly lower than those for 1976-78. End effects in Figure 7 give rise to an implausible trend in numbers of northern common dolphins dur- ing 1975-78. Since 1980, there may have been a decline in this stock, but no smoothed estimates differ signif- icantly. The central stock of common dolphins (Fig. 8) shows evidence of a steep decline from 1977 to 1983, with stability since. The smoothed estimate for 1988 is significantly lower than for all years preceding 1980 (P<0.05), but does not differ significantly from any later estimates. Data on the southern stock of common dolphins are sparse. There may have been a decreas- ing trend (Fig. 9), but unsmoothed estimates fluctuate widely and no smoothed estimates differ significantly. Buckland et al Estimating abundance of tuna-associated dolphin stocks in the eastern tropical Pacific 1975 1976 1977 1978 1979 1980 1981 1982 1983 1984 1985 1986 1987 1983 1989 Figure 8 Smoothed abundance trends of central stock of common dolphin Delphinus delphis in the eastern tropical Pacific. Broken lines indicate approximate 85% confidence limits. See Figure 1 for more details. 1975 1976 1977 1978 1979 1980 1981 1982 1985 1984 1985 1936 1987 1988 1989 Figure 10 Smoothed abundance trends of pooled northern, central, and southern stocks of common dolphin Delphinus delphis in the eastern tropical Pacific. Broken lines indicate approximate 85% confidence limits. See Figure 1 for more details. 1975 1976 1977 1978 1979 1980 1981 1982 1983 1984 1985 1986 1987 1988 1989 Figure 9 Smoothed abundance trends of southern stock of common dolphin Delphinus delphis in the eastern tropical Pacific. Broken lines indicate approximate 85% confidence limits. See Figure 1 for more details. 1975 1976 1977 1978 1979 1980 1981 1982 1983 1984 1985 1986 1987 1988 1989 Figure 1 1 Smoothed abundance trends of northern offshore stock of spotted dolphin Stenella attenuata in the eastern tropical Pacific. Broken lines indicate approximate 85% confidence limits. Estimates and limits were determined from four in- dependent sets of 79 bootstrap replicates, so that the plot indicates uncertainty in the estimates arising from Monte Carlo variation. If data are pooled across stocks of common dolphins (Fig. 10), the 1988 smoothed estimate is significantly lower than all those preceding 1981. Four independent sets of 79 bootstrap replicates were generated for the northern offshore stock of spotted dolphins. The resulting plots, one of which corresponds exactly to Figure 1, are superimposed in Figure 11. If an infinite number of replicates could be carried out for each set, the four plots would be iden- tical. Thus Figure 11 indicates the imcertainty that can be expected in the median and interval estimates due to Monte Carlo variation. Discussion Unsmoothed estimates of relative abundance some- times show larger year-to-year variation than is Fishery Bulletin 90(1). 1992 plausible, even if full allowance is made for the preci- sion of the estimates. An example is the 1983 estimate for the northern offshore stock of spotted dolphins, which is significantly lower than either the 1982 or the 1984 estimate. This has been attributed to the strong El Nino event of that year (Buckland and Anganuzzi 1988a). The change in environmental conditions ap- peared to cause spotted dolphins to split into smaller schools and to disperse more widely than is normal, so that tuna vessels were unable to locate areas of con- centration. If, in normal years when concentrations occur in known areas, there is positive bias in the abun- dance index, then a relatively low estimate might be expected for 1983. This effect would be enhanced if many animals wandered beyond the normal range of the stock, so that the abundance index for 1983 cor- responded to only that portion of the stock remaining within its normal bounds. Such effects may be regarded either as bias that fluctuates over time or as an addi- tional source of variability that is unaccounted for in the variances of the abundance indices. Provided the effects are essentially random, and do not exhibit a con- sistent linear trend over time, the smoothing algorithm described above smooths out the large fluctuations and, in conjunction with the bootstrap, provides variance and interval estimates for the smoothed abundance indices that take full account of variability not allowed for in the variance estimates of the unsmoothed indices. The validity of estimating trends in dolphin abun- dance from tuna-vessel sightings data has been ques- tioned by Edwards and Kleiber (1989). They used a simple simulation model of non-random search vessel effort coupled with clustered distributions of dolphin schools to investigate bias. By allowing the clustering of schools to be slight in one year and extreme in the next, they showed that bias in the relative abundance estimates can be inconsistent between years. They define a change estimate as the ratio of relative abun- dance estimates for the two years. They state, "This two-sample change estimate is only a rough approx- imation to a trend estimate derived from a series of measurements . . . However, conclusions about the ef- fects of inconsistent biases on this change estimate will be valid for trend estimates also, except for the unlikely case in which effects of various inconsistent biases cancel each other out, so that the trend estimate reflects the actual trend, but only fortuitously." (The emphasis on "change" and "trend" is theirs.) They also note that "It is obvious. . .that even relatively small changes of bias can lead to considerably inaccurate estimates of change and, by implication, estimates of trend." If this is so, there would be little value in estimating trends in abundance from tuna-vessel sight- ings data. We question whether the simulation model of Edwards and Kleiber (1989), which is a considerable Table 1 Actual abundance (millions), and expected and simulated | relative-abundance estimate by j ear for a hj-pothetical stock. declining at an annual rate of 5%. Expected abundance is calculated assuming estimates are biased down by 20% in El Nino years (*) and up by 100% in other years Actual Expected Simulated Year abundance estimate estimate 1975 4.00 8.00 8.04 1976* 3.80 3.04 3.37 1977 3.61 7.22 6.86 1978 3.43 6.86 5.86 1979 3.26 6.52 6.87 1980 3.10 6.19 8.66 1981 2.94 5.88 6.26 1982* 2.79 2.23 1.97 1983* 2.65 2.12 3.22 1984 2.52 5.04 4.98 1985 2.39 4.79 5.72 1986 2.28 4.55 4.02 1987* 2.16 1.73 1.65 1988 2.05 4.11 4.01 1989 1.95 3.90 4.75 simplification of reality, allows such strong conclusions. However, we use their results to assess the validity of their argTiments. We take their worst-case scenario of a static environment, using the stratified and smoothed option, and average across their four replicates for the high-density case. The calculations indicate a down- ward bias of about 20% for the "simple, gentle" en- vironmental topography of year 1 and an upward bias of about 100% for the "complex, steep" topography of year 2. Thus, if the population comprised 2500 schools (as in their simulations), the expected estimate would be around 2000 schools in the first year and 5000 in the second, a 2.5-fold estimated increase for a pop- ulation that has constant size. Is this conclusion "valid for trend estimates also"? Suppose a population com- prised 4 million animals in 1975, and decreased at a rate of 5% per annum until 1989. Suppose we again take an extreme scenario in which the "simple, gentle" environmental topography applied in El Nino years, and the "complex, steep" topography applied in all other years. The expectations of the estimates are shown in Table 1. Also shown are simulated estimates, for which errors were generated from a lognormal distribution which yields a coefficient of variation of 15%, close to that observed for estimates based on tuna vessel data. The errors were then added to the ex- pected estimates. The estimated rate of decrease for the expected estimates is 5.0% per annum (SE2.5%), and that for the simulated estimates is 4.7% per annum (SE 2.6%). Thus a scenario of extreme and inconsistent Buckland et al : Estimating abundance of tuna-associated dolphin stocks in the eastern tropical Pacific bias does not invalidate the procedures when applied to a long sequence of estimates. In practice, a rate of change in abundance is unlikely to be roughly constant over such a long time-period, yet tests for trend over a short time-period have low power. Figures 1-10 pro- vide a simple method to test for change over longer time-periods without the necessity of assuming the rate of change is constant. The smoothing procedure used for generating trend estimates can perform poorly at the start (e.g.. Fig. 7) or at the end of a sequence of estimates, so that sharp increases or declines during the first or last year or two should be treated with suspicion. The first and last smoothed estimate in a sequence are especially un- reliable, and are omitted from Figures 1-10. Thus, changes in abundance are assessed relative to 1988 rather than 1989. To assess the current status of dolphin stocks, and the effects of recent levels of mortality, it is necessary to determine whether trends in dolphin abundance are best estimated from tuna vessel data or research vessel data, or whether some combination of estimates from both sources is preferable. Given sufficient data and adequate coverage of the entire range of each stock, research-vessel estimates of trend would be preferred, since they are likely to be less biased. However, Holt and Sexton (1989, 1990ab), to exploit fully the small number of research vessel sightings, made assumptions that might be seriously violated. Firstly, data are pool- ed across all sightings of dolphin schools of at least 15 animals, irrespective of species, to improve precision of effective search-width estimates. This may introduce bias which is not consistent over time, especially if non- target species (those which are seldom associated with tuna, and are therefore seldom encircled by purse seines) have a different effective search width and a different rate of change in abundance than target species. Secondly, although abundance estimates are given by stock, encounter-rate estimates by stock area are ignored for stocks that are not separated in the field. Thus for offshore spotted dolphins, a single abun- dance estimate per year is generated and then prorated by stock area, to yield separate estimates for the north- ern and southern offshore stocks. If the southern off- shore stock became extinct, and the northern offshore stock increased at a rate that ensured overall abun- dance remained constant, the expected trend in re- search vessel estimates would be zero for both stocks. The same applies to common dolphin stocks. The esti- mates of Holt and Sexton indicate that there are large numbers of common dolphins in the western sector of the eastern tropical Pacific, yet the species is seldom recorded there. Using the estimation methods of Holt and Sexton, valid trend estimates from research vessel data are not available separately for northern and southern offshore stocks of spotted dolphin or for the main stocks of common dolphin. In Figures 12-15 we show the valid estimates of trend (i.e., those obtained after pooling data from stocks that are not differentiable in the field) from the research-vessel relative abundance estimates for 1986-89, taken from Sexton et al. (1991) and Gerro- dette and Wade (1991). Also shown are the corre- sponding unsmoothed trend estimates from tuna vessel data. Vertical bars show ± 2 standard errors. Plots are based on the relative abundance estimates and stan- dard errors of Tables 2 and 3. The research vessel estimates indicate changes in abundance that are biologically implausible, even with full allowance for the estimated precision of the estimates. Thus either the precision of the surveys is appreciably worse than estimated or there is strong and inconsistent bias in the estimates from one year to the next. By contrast, despite the concerns over the validity of tuna vessel estimates, they yield biologically plausible rates of change during 1986-89 when the precision of the estimates is accounted for. 5 c o = 4 5 T T T T Abundance ■ ^ • Relative • 0 86 87 88 89 Year Figure 12 Unsmoothed abundance trends of northern and southern off- shore stocks of spotted dolphin Stenella atteniiata in the eastern tropical Pacific, estimated from research (solid line) and tuna vessel data. Vertical bars are ± 2 standard errors. Fishery Bulletin 90|l). 1992 Year Figure 13 Unsraoothed abundance trends of eastern stock of spinner dolphin Stenella longirostris in the eastern tropical Pacific, estimated from research (solid line) and tuna vessel data. Ver- tical bars are ±2 standard errors. IX 8 o c (0 ■□ c < > ^ 2 0) a: Figure 15 Unsmoothed abundance trends of northern, central, and southern stocks of common dolphin Delphinus delphis in the eastern tropical Pacific, estimated from research (solid line) and tuna vessel data. Vertical bars are ± 2 standard errors. 1.5 (Millions) fo T r r Abundance o p T \ 01 > 'i 0.3 lU ■ ■ a: 0 86 87 88 89 Year Figure 14 Unsmoothed abundance trends of whitebelly stock of spinner dolphin Stenella longirostris in the eastern tropical Pacific, estimated from research (solid line) and tuna vessel data. Ver- tical bars are ±2 standard errors. Acknowledgments We are grateful to Dr. J. Joseph, Dr. M. Hall, and Dr. M. Scott for comments on the methods outlined here, and to two reviewers and Dr. L. Jones for their constructive comments and criticisms. We also ac- knowledge the recent and continuing efforts of the Southwest Fisheries Science Center to evaluate methods for analyzing tuna vessel and research vessel sightings data; their program of work forced us to address more carefully the issue of how to estimate and test for trends in abundance. Citations Anganuzzi, A. A., and S.T. Buckland 1989 Reducing bias in estimated trends from dolphin abun- dance indices derived from tuna vessel data. Rep. Int. Whal- ing Comm. 39:323-334. Anganuzzi, A. A., S.T. Buckland. and K.L. Cattanach 1991 Relative abundance of dolphins associated with tuna in the eastern tropical Pacific, estimated from tuna vessel sight- ings data for 1988 and 1989. Rep. Int. Whaling Comm. 41: 497-506. Au, D., W.L. Perryman, and W. Perrin 1979 Dolphin distribution and the relationship to environmental features in the eastern tropical Pacific. Admin. Rep. LJ-79-43, Southwest Fish. Sci. Cent., NMFS, NOAA, La Jolla, CA 92038, 59 p. Buckland et al.: Estimating abundance of tuna-associated dolphin stocks in the eastern tropical Pacific 1 I Table 2 Unsmoothed relative-abundance estimates (standard errors in parentheses) of some eastern tropical Pacific, calculated from research vessel data collected 1986-89. stocks of dolphin in the Offshore Eastern Year spotted dolphin spinner dolphin Whitebelly spinner dolphin Common dolphin 1986 1527 (261)** 716 1987 2388 (377) 707 1988 2549 (476) 902 1989 3560 (634)** 1200 (152) (138) (191) (254) 657 (140) 750 (159) 821 (174) 759 (248) 1810 (437)* 1026 (298)tTT 5263 (1368)*TT 2586 (587)t •Estimates differ significantly (P<0.05) t Estimates differ significantly (P<0.05) ••Estimates differ significantly (P<0.01) TT Estimates differ significantly (P<0.01) Table 3 Unsmoothed relative-abundance estimates (standard errors in parentheses) of some stocks of dolphin in the | eastern tropical Pacific calculated from tuna vessel data collected 1986-89. Offshore Eastern Whitebelly Common Year spotted dolphin spinner dolphin spinner dolphin dolphin 1986 3484 (342) 590 (118) 595 (119) 532 (159) 1987 3627 (420) 363 (100) 937 (170) 271 (132) 1988 3048 (439) 665' (119) 575 (109) 487 (167) 1989 3640 (337) ficantly (P<0.05) 381* (74) 748 (105) 408 (111) * Estimates differ signi Buckland, S.T. 1984 Monte Carlo confidence intervals. Biometrics 40: 811-817. Buckland, S.T., and A. A. Anganuzzi 1988a Trends in abundance of dolphins associated with tuna in the eastern tropical Pacific. Rep. Int. Whaling Comm. 38: 411-437. 1988b Comparison of smearing methods in the analysis of minke sightings data from IWC/IDCR Antarctic cruises. Rep. Int. Whaling Comm. 38:257-263. Burnham, K.P., D.R. Anderson, and J.L. Laake 1980 Estimation of density from line transect sampling of biological populations. Wildl. Monogr. 72, 202 p. Cooke, J.G. 1985 On the relationship between catch per unit effort and whale abundance. Rep. Int. WhaUng Comm. 35:511-519. Edwards, E.F., and P.M. Kleiber 1989 Effects of nonrandomness on line transect estimates of dolphin school abundance. Fish. Bull., U.S. 87:859-876. Gerrodette, T., and P.R. Wade 1991 Monitoring trends in dolphin abundance in the eastern tropical Pacific using research vessels over a long sampling period: Analysis of 1989 data. Rep. Int. Whaling Comm. 41: 511-515. Hall, M.A., and S.D. Beyer 1988 Incidental mortality of dolphins in the eastern tropical Pacific. Rep. Int. Whaling Comm. 38:439-441. Hammond. P.S.. and J.L. Laake 1983 Trends in estimates of abundance of dolphins (SteneUa spp. and Delphinus delphis) involved in the purse-seine fishery for tunas in the eastern Pacific Ocean, 1977-81. Rep. Int. Whaling Comm. 33:565-588. Holt, R.S. 1985 Estimates of abundance of dolphin stocks taken inciden- tally in the eastern tropical Pacific yellowfin tuna fishery. Admin. Rep. LJ-85-20, Southwest Fish. Sci. Cent., NMFS, NOAA, La JoUa, CA 92038, 32 p. 1987 Estimating density of dolphin schools in the eastern tropical Pacific Ocean by line transect methods. Fish. Bull., U.S. 85:419-434. Holt, R.S., and J.E. Powers 1982 Abundance estimation of dolphin stocks involved in the eastern tropical Pacific yellowfin tuna fishery determined from aerial and ship surveys to 1979. Tech. Memo. 23, Southwest Fish. Sci. Cent., NMFS, NOAA, La Jolla, CA 92038, 95 p. Holt, R.S., and S.N. Sexton 1989 Monitoring trends in dolphin abundance in the eastern tropical Pacific using research vessels over a long sampling period: Analyses of 1987 data. Rep. Int. Whaling Comm. 39:347-351. 1990a Monitoring trends in dolphin abundance in the eastern tropical Pacific using research vessels over a long sampling period: Analyses of 1986 data, the first year. Fish. Bull., U.S. 88:105-111. ,2 Fishery Bulletin 90(1). 1992 1990b Monitoring trends in dolphin abundance in the eastern tropical Pacific using research vessels over a long sampling period: Analyses of 1988 data. Rep. Int. Whaling Comm. 40:471-476. Perrin, W.F., P.A. Akin, and J.V. Kashiwada 1991 Geographic variation in external morphology of the spin- ner dolphin Stenella longirostris in the eastern Pacific and im- plications for conservation. Fish. Bull., U.S. 89:411-428. Polacheck, T. 1987 Relative abundance, distribution and inter-specific rela- tionship of cetacean schools in the eastern tropical Pacific. Mar. Mammal Sci. 3:54-77. Sexton, S.N., R.S. Holt, and D. DeMaster 1991 Investigating parameters affecting relative estimates in dolphin abundance in the eastern tropical Pacific from research vessel surveys in 1986, 1987, and 1988. Rep. Int. Whaling Comm. 41:517-524. Smith, K.L. 1988 Calibration and smoothing of relative dolphin abundance estimates. MSc. diss., University of Strathclyde. Smith, T.D. 1983 Changes in sizes of three dolphin {Sfenella spp.) popula- tions in the eastern tropical Pacific. Fish. Bull., U.S. 81:1-14. Velleman, P.F.. and D.C. Hoaglin 1981 Applications, basics and computing of exploratory data analysis. Duxbury Press, Boston. Abstract.- Osteological differ- ences confirm the validity of two spe- cies of Grammatorcynus, G. bicari- natus (Quoy and Gaimard 1825) and the long-recognized G. bilineattis (Riippell 1836). In addition to having fewer gill rakers (12-15 vs. 18-24), a smaller eye (3.1-4.6% vs. 4.0-6.0% FL), small black spots on the lower sides of the body, and reaching a larger size (110cm FL vs. 60cm), G. bicarinatus differs from G. biline- atus in having a shorter neurocra- nium, shorter parasphenoid flanges, lower posterior edge of maxillary shank, shorter quadrate process, narrower first postcleithrum, wider ethmoid, wider vomer, wider lach- rymal, longer teeth, wider palatine tooth patch, wider opercle, and a thin posttemporal shelf between the anterior processes. All but one of the 16 osteological differences previous- ly found between Grammatorcynus bilineatus and Scomberomorus and Acanthocybium are confirmed with the inclusion of G. bicarinatus in the genus. Grammatorcynus bilineatus is widespread in tropical and sub- tropical waters of the Indo-West Pacific from the Red Sea to Tokelau Islands in Oceania. The range of G. bicarinatus is restricted to the west- ern and eastern coasts of Australia and southern Papua New Guinea. Morphology, systematics, and biology of the double-lined mackerels [Grammatorcynus, Scombrldae) Bruce B. Collette Systematics Laboratory. National Marine Fisheries Service, NOAA National Museum of Natural History, Washington, DC 20560 Gary B. Gillis Observer Program, Alaska Fisheries Science Center, National Marine Fisheries Service NOAA, 7600 Sand Point Way NE, Seattle, Washington 981 15-0070 Current address: Department of Ecology and Evolutionary Biology University of California, Irvine, California 92715 Manuscript accepted 18 December 1991. Fishery Bulletin, U.S. 90:13-53 (1992). Until recently, most authors consid- ered the genus Grammatorcynus to be monotypic (Fraser-Brunner 1950, Silas 1963, Zharov 1967, Collette 1979). Electrophoretic work (Lewis 1981, Shaklee 1983) indicated there were two species of double-lined mack- erels in Australia. This was confirmed by Collette (1983) who showed there are two species: the double-lined mackerel or scad G. bilineatus, (Riip- pell 1836), widespread in the Indo- West Pacific, with more gill rakers (18-24), a larger eye (4.0-6.0% FL), and a smaller maximum size (60 cm FL); and the shark mackerel G. bica- rinatus (Quoy and Gaimard 1825), restricted to the waters of northern Australia and southern New Guinea, with fewer gill rakers (12-15), a smaller eye (3.1-4.6% FL), and a larger maximum size (110 cm FL). All morphological information concern- ing Grafnimatorcynus in Collette (1979) and Collette and Russo (1985b) was based solely on G. bilineatus. The purposes of this paper are to describe osteological differences be- tween the two species of Gramma- torcynus, redefine the genus and both species, and summarize the literature on both species. The paper is divided into two parts. Part 1, Comparative Morphology, contains descriptions and illustrations of mor- phometry, meristic characters, soft anatomy, and osteology of the two species of Grammatorcynus; com- parisons are made with Scombero- morus and Acanthocybium- where appropriate. Part 2, Systematics and Biology, contains a generic descrip- tion and accounts of both species, in- cluding synonymy, types of nominal species, diagnoses (based on char- acters from the first section), size, biology, interest to fisheries, geo- graphic distribution, and material examined. Methods and materials Methods are those used by Collette and Russo (1985b) in a revision of Scomberomorus, and by Collette and Chao (1975) in a revision of the bonitos (Sardini). Material of Grammatorcynus is listed at the end of each species ac- count; 80 specimens of G. bilineatus and 11 G. bicarinatus. Abbreviations of institutions housing the material follow Leviton et al. (1985). Com- parative material oi Scomberomorus and Acanthocybium was listed in the species accounts in Collette and Russo (1985b). 13 14 Fishery Bulletin 90(1). 1992 '^"•iirairi'^i-itlliiillftlM'i"'^- B Figure I Species of Gr animator cynus. (A) G. bilineatus (from Evermann and Seale 1907. fig. 3, holotype oi Nesogrammus piersoni, 372mm FL, Philippine Is.); (B) G. bicarinatus (from McCulloch, 1915, p. 1, fig. 1, 925 mm FL, New South Wales, Australia). Part 1: Comparative morphology Morphological characters useful for distinguishing be- tween species of Grammatorcynus and for evaluating phylogenetic relationships of the genus are divided into six categories: lateral line, color pattern, morphometry, meristic characters, soft anatomy, and osteology. Lateral line The genus Grammatorcynus differs from all other genera of Scombridae in having two lateral lines, hence their common name, double-lined mackerels. The dorsal-most lateral line is slightly convex, originates near the dorsal portion of the opercle, and continues posteriorly until it converges with the second lateral line, just anterior to the median caudal keel. The sec- ond lateral line originates from the first at a point below the first four spines of the dorsal fin. It starts ventrally, running under, or just posterior to, the pec- toral fin, and abruptly turns into a concave line that continues posteriorly until meeting the dorsal lateral line (Fig. 1). The function of this additional lateral line is unknown. The characteristic two lateral lines are discernible in specimens as small as 56.9mm SL (Nishikawa 1979:133). Anomalies in the pattern of the lateral lines are occasionally found, but none appear to be species specific (Fig. 2; Silas 1963: fig. 3). Color pattern Dark spots are usually found on the ventral portion of G. bicarinatus (Fig. IB). The spots are smaller than the pupil, originate near the ventral border of the oper- Collette and Gillis Osteological differences between two species of Grammatorcynus 15 Figure 2 Variations in lateral line pattern in Grammatorcynus. (a) Usual pattern in G. bilineatus; (b-d) variations in pattern in G. bilineatus; (b) Australia, 410mm FL; (c) Queensland, ^16mm FL; (d) Queensland, 400mm FL; (e) usual pattern in G. bicarinatus; (f) variation in pattern in G. hicarinatus. Western Australia, 765 mm FL. 28 22 ORBF, mm 0^ 'A 10 / / " ) 1 1 1 1 15 46 77 ' 108 139 170 HDL, mm Figure 3 Orbit length (ORBF) compared with head length (HDL) in Grammatorcynus. Open circles = G. bilin£atus. squares = G. bicarinatus. shows the range and mean of all the characters as thousandths of fork length, and eight of the characters as thousandths of head length (Table 1). Scatter diagrams, with regression lines, show two of the best morphometric characters: G. hicarinatus has a smaller orbit (Fig. 3), and a longer first dorsal fin base (Fig. 4). MerJstJc characters Numbers of fin rays (first dorsal spines, second dorsal rays, dorsal fmlets, anal rays, anal finlets, and pectoral rays), gill rakers, and teeth on the upper and lower jaws are systematically valuable in Grammatorcynus. They are discussed in the relevant osteological sections of the paper. culum, and continue posteriorly to the anal fin. They are found below the ventral lateral line on both sides of the fish. No spots were present in the two smallest specimens examined (AMS IB.5207-8, 306-315mm FL). Spots are never present in G. bilineatus (Fig. lA). Morphometric characters In addition to fork length, 26 measurements were routinely made on all specimens. Several morphometric characters separate the two species. A summary table Soft anatomy Viscera Emphasis was placed on the appearance of the viscera in ventral view, after removal of an oval segment of the belly wall. Previous descriptions of the viscera of Grammatorcynus include Kishinouye (1923), Silas (1963), and Collette and Russo (1985b). The anterior end of the liver abuts the transverse septum anteriorly in the body cavity. The liver has three lobes. The right and left lobe are longer than the middle lobe, with the right lobe being longest (Fig. 5c-d). The liver is similar in shape in Scomberomorus, 16 Fishery Bulletin 90(1). 1 992 rable 1 Morphometric comparison of Grammatorcynus bilineatus and G. bicarinatus. Character G . bicarinatus G bilineatus N Min Max Mean SD N Min Max Mean SD Fork len^h (thousandths) 10 306 825 551 186 64 226 575 408 77 Snout-A 7 596 626 613 10 61 581 641 606 13 Snout-2D 7 536 558 549 8 61 528 619 547 14 Snout- ID 9 267 301 280 11 64 276 322 295 9 Snout-P2 9 234 272 253 13 63 236 306 258 12 Snout-Pl 9 197 230 216 10 63 199 245 226 9 P1-P2 10 91 255 115 49 62 90 135 101 7 Head length 10 191 223 207 9 64 197 236 218 7 Max. body depth 8 177 210 192 13 57 164 234 196 14 Max. body width 8 105 129 115 8 56 91 136 114 9 PI length 10 118 137 127 5 63 106 142 126 8 P2 length 10 65 81 74 5 63 70 87 77 3 P2 insertion-vent 7 313 345 332 12 62 262 354 328 14 P2 tip-vent 9 238 281 260 15 61 228 275 251 10 Base ID 9 253 272 264 6 63 207 261 235 11 Height 2D 6 97 111 103 5 54 82 116 98 7 Base 2D 10 76 102 90 8 62 68 118 102 9 Height A 10 94 116 104 8 49 67 114 94 9 Base A 9 66 91 80 8 63 73 105 87 7 Snout (fleshy) 10 77 88 81 4 64 58 90 80 5 Snout (bony) 10 64 76 70 4 64 60 80 72 5 Maxilla length 10 91 110 102 6 63 89 108 98 5 Postorbital 10 87 98 92 3 62 78 98 91 3 Orbit (fleshy) 10 31 46 37 5 64 40 60 49 4 Orbit (bony) 10 48 69 59 8 64 53 88 68 6 Interorbital 9 59 71 64 4 62 56 74 62 3 2D-caudal 9 412 475 454 27 60 427 496 470 13 Head length (thousandths) 11 64 165 112 33 64 50 126 89 17 Snout (fleshy) 11 379 410 393 8 64 248 397 366 21 Snout (bony) 11 313 356 340 16 64 281 357 329 16 Maxilla length 11 475 510 495 12 63 420 480 448 15 Postorbital 11 412 471 446 17 62 350 450 419 15 Orbit (fleshy) 11 164 211 179 16 64 191 257 226 14 Orbit (bony) 11 238 319 282 25 64 252 381 313 24 Interorbit 10 274 322 308 13 62 253 327 283 13 but in Acanthoeybium the right and left lobes are about the same size. Two efferent vessels lead directly from the anterior surface of the liver into the sinus venosus. The stomach is sometimes visible in ventral view, partially covered by the liver and caecal mass, but often completely hidden. Stomach contents included crusta- ceans and small fishes. The pyloric portion of the intestine arises from the anterior end of the stomach, where the main branches of the pyloric caeca join the intestine. The caeca branch and form a dense dendritic conglomeration, the caecal mass. The intestine continues posteriorly as a simple straight tube to the anus. A straight intestine is also found in Acanthoeybium (Fig. 5b) and S. niphonius, but all other species of Scomberomorus have folds (2 or 4) in the intestine (Fig. 5a). Osteology The osteological description is divided into five sections: skull, axial skeleton, dorsal and anal fins, pectoral girdle, and pelvic girdle. Osteological terminology and organization generally follow that of Collette and Russo (1985b). Skull Description of the skull is presented in two sec- tions: neurocranium (Figs. 6-9) and branchiocranium. Neurocranium Following a general description of the neurocranium, the four major regions are dis- cussed: ethmoid, orbital, otic, and basicranial. General characteristics In dorsal view (Fig. 6), the neurocranium of Grammatorcynus is more or less triangular in shape, narrow at its anterior margin. Collette and Gillis: Osteological differences between two species of Grammatorcynus 17 205 200 290 380 470 560 o\ : : FL , mm Figure 4 Length of first dorsal fin base (BID) compared with forl< length (FL) in Grammatorcynus. Open circles = G. bilineatus, squares = G. bicarinatus. widening posteriorly. It is intermediate in shape be- tween the elongate neurocranium of Acanthocyhium, Scomber, and Rastrelliger , and the shorter, wider neurocranium of Thunnus. The posterodorsal surface is marked by a median ridge (supraoccipital crest), with two parallel ridges on either side. These five thin ridges of bone form six grooves, three on each side: dilator (very shallow), temporal (quite deep), and supratem- poral (most easily seen in lateral view) (Allis 1903:49). The median ridge originates just posterior to the thin, oval pineal foramen located between the posterior, median edges of the frontal bones. This ridge becomes larger posteriorly, and forms the supraoccipital crest. Internal or temporal ridges originate at the posterior portion of the frontals (midlevel of the orbit), continu- ing posteriorly to the epiotic. External or pterotic ridges also originate near the posterior margin of the frontals, continuing posteriorly to the pterotic. Neurocrania of the two species of Grammatorcynus differ in size, relative to fork length. Length of the neurocranium, measured from the anterior tip of the vomer to the posterior margin of the basioccipital, is slightly longer in G. bilineatus (14-16% FL) than in G. bicarinatus (13% FL). Ethmoid region This region is composed of the ethmoid, lateral ethmoid, and vomer. The nasal bone lies lateral to the ethmoid and lateral ethmoid, and, therefore, is included here. Ethmoid The ethmoid (dermethmoid) has a smooth flat dorsal surface that is partially overlapped by the frontals. It connects ventrally to the vomer, posteriorly to the lateral ethmoids, and anterolateral- ly to the nasals. Its anterior border is nearly straight, with an anteromedian projection, unlike the relatively smooth, concave border in Scomberomorus and Acan- thocybium. The ethmoid is clearly visible in dorsal view (Fig. 6), and is wider, relative to the length of the neurocranium, in G. bicarinatus (width 25-28% of length) than in G. bilineatus (19-21%). Lateral ethmoid The lateral ethmoids (pareth- moids) are massive, paired bones that extend down- ward from the middle region of the frontals and form the anterior margin of the orbit and the posterior and mesial walls of the nasal cavity. The ventral surface of the lateral ethmoid bears an articulating surface for the palatine, and the posterolateral process serves as an articulation surface for the lachrymal. The lateral expansion of the bone is greater in G. bicarinatus (45-50% of neurocranium length) than in G. bilineatus (39-42%) (Fig. 8). Vomer The anterior process of the vomer bears a circular or oval patch of fine teeth on its ventral sur- face. Its pointed posterior end is firmly ankylosed dor- sally with the parasphenoid. The anterior process is wider in G. bicarinatus (16-18% of neurocranium length) than in G. bilineatus (13-15%) (Fig. 8). IMasal The nasal bones are flat, elongate bones that articulate with the lateral edge of the frontals. They project out beyond the ethmoid and, from a dor- sal view, reach about as far anteriorly as the vomer. There is no such projection of the nasal bones in Scomberomorus or Acanthocybium. Length divided by width is 2.8-3.4 in Grammatorcynus, which is inter- mediate between the ranges oi Scomberomorus (2.0- 3.1) and Acayithocybiuyn (3.1-4.2). The anterior end of the bone forms a short, slightly angled arm. No differences were found between the nasals of the two species of Grammatorcynus. Orbital region The orbit (Fig. 7) is surrounded by the posterior wall of the lateral ethmoid, the ven- tral side of the frontal, the pterosphenoid, sphenotic, prootic, suborbital, and lachrymal bones. The left and right orbits are partially separated by the basisphenoid. The sclerotic bones enclose the eyeballs. The orbit of G. bilineatus is larger than that of G. bicarinatus (Fig. 7), reflecting the difference in orbit length (Fig. 3). The maximum height of the orbit measured from the parasphenoid to the pterosphenoid is 24-25% of neurocranium length in G. bilineatus vs. 16-17% in G. bicarinatus. Orbit length in G. bilineatus is 51-54% of neurocranium length vs. 47-49% in G. bicarinatus. Fishery Bulletin 90(1), 1992 9 C LIVER ^:::::>::-M CAECAL MASS • • • • ^ ■ • • • • • • • • i I NTESTI NE GONAD P?aS?X^ t^i: STOMACH GALL BLADDER ^ U RINARY BLADDER GAS BLADDER Figure 5 Ventral view of viscera, (a) Scomberomorus tnaculatus, Georgia, 290mm FL; (h) Acan- thocybium solandri, Campeche Banks, Mex- ico, 1280mm FL; (c) Grammatorcynus bilineatus, Marshall Is.. 424 mm FL; (d) G. bicarinatus, Australia. Frontal The paired frontals form the largest portion of the dorsal surface of the neurocranium. A small, elongate oval pineal opening is present between the posterior ends of the frontals. A larger and more irregular foramen is present in Acanthocybium, but Scomberomorus lacks this opening (Collette and Russo 1985b:figs. 11-12). In Scomberomorus and Acanthocybium, the frontals form a median ridge that increases in height posteriorly and joins the supraocoipital crest. Grammatorcynus lacks this ridge and the supraoccipital crest begins posterior to the pineal opening, giving the top of the skull a much flatter appearance than in the other two genera. In ventral view (Fig. 8), the left and right frontals articulate with the pterosphenoids at the anterior end of a median opening into the brain cavity. The ridge around the anterior end of this space forms a point and Collette and Gillis Osteological differences between two species of Grammatorcynus 19 SPHENOTIC PTEROTIC FRONTAL NASAL INTERCALAR EPIOTIC EXOCCIPITAL VOMER ETHMOID _ LATERAL ETHMOID a FIRST VERTEBRA SUPRAOCCIPITAL PARIETAL Figure 6 Dorsal view of skulls in Grammatorcynus. (a) G. bilirwatiis, Scott Reef, Timor Sea, 453mm FL; (b) G. bicarinatus. Western Australia, Exmouth Gulf, 765 mm FL. extends almost to the ethmoid in G. bilineatus. The ridge curves around the anterior end of the space and ends distinctly more posteriorly in G. bicarinatus. This difference cannot be seen in the ventral view of the skulls (Fig. 8) because the median part of the opening is obscured by the parasphenoid, so a separate outline figure has been made (Fig. 9). Pterosphenoid The pterosphenoids (alisphe- noids) form the posterodorsal margin of the orbit. They serve as the base for the median basisphenoid, and abut the prootics posteriorly and the frontals and sphenotics laterally. 20 Fishery Bulletin 90(1). 1992 SUPRAOCCIPITAL CREST PTEROSPHENOID FRONTAL ETHMOID PTEROTIC LATERAL ETHMOID a FIRST VERTEBRA EXOCCIPITAL PROOTIC BASISPHENOID Figure 7 Lateral view of skulls in Grammatorcynus. (a) G. hUineatus, Scott Reef, Timor Sea, 453 mm FL; (b) G. hkariiiatus. Western Australia, Exmouth Gulf, 765mm FL. Sclerotic The sclerotic bones consist of two thickened, semicircular segments connected by carti- lage on the inner surface and by corneal membranes on the outside. The sclerotic bones of Grammatorcynus are relatively larger and thinner compared with Scom- beromorus and Acanthocybium. Basisphenoid The basisphenoid is a small, median, Y-shaped bone that connects the prootics and pterosphenoids dorsally with the parasphenoid ventral- ly (Fig. 7). The dorsal compressed vertical base bears a slight anterior process, but no posterior process. This is similar to the condition in Scomber omorus, but the anterior process is much shorter in Grammatorcynus. The basisphenoid is longer in G. bilineatus since the height of the orbit is greater in this species compared with G. bicarinatus. Infraorbitals The bones of the infraorbital series (Fig. 10) enclose the infraorbital branch of the lateral sensory canal system. The canal enters the infraorbital series at what is usually considered the last element (dermosphenotic), and continues around the orbit, terminating on the first infraorbital (lachrymal). The lachrymal, the first and largest element, is elongate with a mesially-directed articular process just anterior to the middle of the bone. It covers part of the maxilla, and articulates with the lateral ethmoid Collette and Gillis. Osteological differences between two species of Grammatorcynus SPHENOTIC FRONTAL LATERAL ETHMOID PROOTIC V — INTERCALAR BASIOCCIPITAL VOMER a PARASPHENOID FIRST VERTEBRA EXOCCIPITAL PTEROSPHENOID PTEROTIG Figure 8 Ventral view of skulls in Grammatorcynus. (a) G. hilineatus, Scott Reef, Timor Sea, 453mm FL; (b) G. bicarinatus. Western Australia, Exmouth Gulf, 765 mm FL. dorsally by the articular process. The process is larger in G. bicarinatus, making the lachrymal wider (30-35% of total bone length) than in G. hilineatus (27-30%). The anterior portion has a small notch in it, much more indistinct than the forked anterior region in Scorn- beromorus (Fig. 10a). The posterior region is distinct- ly forked, with the ventral arm being wider and longer than the dorsal arm. The second infraorbital connects to the forked pos- terior region of the lachrymal. It is a small, elongate bone. The third infraorbital is an elongate, tubular bone that connects to the posterior portion of the second 22 Fishery Bulletin 90(1). 1992 a Figure 9 Outline of pterosphenoid opening on ventral side of skull in Grammatorcynus. (a) G. bilineatus, Scott Reef, Timor Sea, 453 mm FL; (b) G. bicarinatus, Western Australia, Exmouth Gulf, 765mm FL. infraorbital. It has a large, mesial, shelflike extension (subocular shelf of Smith and Bailey 1962). The fourth through penultimate elements total 13 in a specimen of G. bilineatus (Fig. 10c), are small, and are easily lost with cheek scales during dissection. No special effort was made to compare these bones in the two species. Otic region This region encloses the otic cham- ber inside the skull, and is formed by the parietal, epiotic, supraoccipital, prootic, pterotic, sphenotic, and intercalar (opisthotic) bones. Parietals The parietals articulate with the frontals anteriorly, the supraoccipital mesially, the pterotics laterally, sphenotics ventrally, and epiotics posteriorly. There is a short inner lateral crest on the parietals and epiotics, but this crest does not originate on the frontals as it does in Scomberomorus and Acanthocybium. Epiotics The epiotics are irregular bones bounded by the parietals anteriorly, the supraoccipital mesially, the exoccipitals posteriorly, and the pterotics laterally. The medial process of the posttemporal bone attaches to a distinct roughened process on the posterior corner of the epiotic. Scomberomorus has a roughened area at the posterior end of the fronto- epiotic crest rather than a distinct process. Supraoccipital The supraoccipital forms the dorsomedian portion of the posterior end of the neuro- cranium. It bears a well-developed crest that continues forward onto the parietals but stops at the pineal opening instead of extending all the way forward onto the frontals as in Scomberomorus. The supraoccipital consists of a thin crest on a roughly hexagonal base. The crest extends down over the exoccipitals along the median line where the dorsal walls of the exoccipitals suture with each other. It extends posteriorly over the first vertebral centrum (Fig. 7). Prootics In ventral view (Fig. 8), the prootics connect with all the bones in the posterior part of the neurocranium. Each prootic is bordered ventrally by the parasphenoid; posteriorly by the basioccipital, ex- occipital, and intercalar; laterally by the pterotic and sphenotic; and anteriorly by the parasphenoid and basisphenoid. The prootics are irregular in shape and meet each other along the ventromedian line of the brain case to form the posterior portion of the myodome. Pterotics The pterotics form the lateral pos- terior corners of the neurocranium. Each pterotic is produced posteriorly to form a spine. A pterotic ridge continues anteriorly onto the parietal, but does not extend onto the posterior part of the frontal as it does in Scomber om^orus. In ventral view (Fig. 8), the pterotics articulate with the sphenotics anteriorly and the prootics and intercalars medially. Sphenotics The sphenotics form the most pos- terior dorsolateral part of the roof of the orbit. They continue the outer lateral shelf from the frontals, and articulate with the pterosphenoid medially and the prootic and pterotic posteriorly. A fossa at the junc- ture of the sphenotic and pterotic receives the anterior condyle of the hyomandibula. In dorsal or ventral view, the distance between the tips of the two sphenotics is the widest portion of the cranium, 60-67% the length of the neurocranium in Grammatorcynus. Intercalars The intercalars (opisthotics) are flat bones that form part of the posterior border of the neurocranium interposed between the pterotics and ex- occipitals. The anterior portion on the dorsal surface is concealed by the overlapping pterotic, thus expos- ing the bone on the dorsal surface less than on the ventral surface (compare in Figures 6 and 8). Each intercalar has a roughened area on its dorsal surface to receive the lateral arm of the posttemporal. There is no posterior projection from the intercalars in Gram- matorcynus or Acanthocybium as there is in eight species oi Scomberomonis, such as S. commerson and S. concolor (Collette and Russo 1985b: figs. 11a and 12b). Basicranlal region This region consists of the parasphenoid, basioccipital, and exoccipital bones, and forms the posteroventral base of the skull. Parasphenoid The parasphenoid is a long, cross-shaped bone. It articulates with the vomer ante- riorly and forms the ventral axis of the skull. It also Collette and Gillis Osteological differences between two species of Crammatorcynus 23 dennosphenotic anterior process cheek scales Figure 10 Left infraorbital bones in lateral view, (a) Scomberomorus maculatus, Cape Hatteras. NC, 534 mm FL; (b) Acanthocybium solandri, Revillagigedos Is., 1068mm FL; (c) Grarmnatocrynus bilineatus, Timor Sea, 453mm FL. forms the ventral border of the orbits and connects with the lateral ethmoids, basisphenoid, prootics, and basioccipital bones dorsally. The lateral wings of the parasphenoid extend dorsolaterally along the ventral ridge of the prootic bones on either side, and have pointed ends which form part of the anteroventral wall of the posterior myodome. Posteriorly, the parasphe- noid bifurcates into two lateral flanges that attach dorsally to the corresponding posteroventral flanges of the basioccipital bone, and surround the posterior 24 Fishery Bulletin 90(1). 1992 opening of the posterior myodome. These flanges are longer in G. bilineatus (18-21% of neurocranium length) than in G. bicarinatus (14%), making the posterior opening of the posterior myodome larger in G. bilineatus (Fig. 8). A ventrally projecting median keel is present in the area anterior to the origin of the lateral flanges. In ventral view, the parasphenoid nar- rows posteriorly until near the region of the median keel, where it widens slightly before the lateral wings. The anterior portion and the region just anterior to the lateral wings are about equal in width. In Gramma- torcynus, the shaft of the parasphenoid is narrower than that of Scomberomorus and Acanthocybium. In G. bilineatMS, the contour of the parasphenoid is con- cave, making the orbit larger than in G. bicarinatus, in which the parasphenoid is flat (Fig. 7). BasioccJpital The basioccipital has lateral flanges on either side of the skull and forms the roof and lateral walls of the posterior myodome. The lateral flanges expand ventrally to meet the flat posterior flanges of the parasphenoid. Anteriorly, the basi- occipital is attached to the prootics and dorsally with the exoccipitals. The first vertebral centrum attaches to the posterior surface of the basioccipital. Exoccipital The exoccipitals connect the skull with the first vertebra dorsally. The exoccipital artic- ulates with the epiotic and supraoccipital bones antero- dorsally, the intercalars laterally, and with the other exoccipital posterodorsally. In ventral view, the ex- occipital articulates with the prootic anteriorly, basi- occipital ventromedially, and intercalar laterally. In posterior view, the foramen magnum is framed by the exoccipitals. Branchiocranium The branchiocranium is divided into five sections: mandibular arch, palatine arch, hyoid arch, opercular apparatus, and branchial apparatus. Mandibular arch The man- dibular arch is composed of the upper jaw (premaxilla, maxilla, and supramaxilla) and the lower jaw (dentary, angular, and retro- articular). Teeth are borne on the premaxilla and dentary, and the number of teeth on these bones differs between species. Dentition Long, thin, slightly laterally compressed teeth are present in a single row in the upper and lower jaws of Grammatorcynus. Scomberomo- rus has large, triangular, lateral- ly compressed teeth similar to those of Acanthocybium, which are blunter and more tightly compressed. The length of the jaw teeth differs between the species: G. bica- rinatus has longer teeth than G. bilineatus (maximum length 6% vs. 4% dentary length). The number of jaw teeth in Grammatorcynus also varies. Teeth are often broken or lost, so the range in mean tooth count may not reflect accurately the actual number of teeth. However, the maximum number of teeth is useful. Grammatorcynus bicarinatus has a lower maximum tooth count on its upper jaw than G. bilineatus (25 vs. 37), and the same is true of the lower jaw, (23 vs. 32; Table 2). The maximum number of jaw teeth present in Scomberomorus is slightly higher than G. bilineatus (39, range 5-39 in the upper jaw; and 37, range 4-37 in the lower jaw). Collette and Russo (1985b) noted that in Scomberomorus, the species with the fewest teeth has the fewest gill rakers and the species with the most teeth has the most gill rakers. There is a similar cor- relation in Grammatorcynus: G. bilineatus also has more gill rakers (18-24 vs. 12-15 in G. bicarinatus). Premaxilla The premaxilla (Fig. 11) is a long, curved bone with an arrowhead-shaped anterior end that extends dorsally and posteriorly as an ascending process. The posterior shank of the premaxilla is elongate and bears a row of 14-37 long, thin teeth on its ventral margin. There are two articular facets for the overlying maxilla at the junction of the posterior margin of the ascending process with the shank por- tion. Ascending processes of both premaxillae are closely approximated to each other mesially and fit into the median groove of the ethmoid bone. The ascend- ing process forms an angle of 55-67° with the shank: G. bilineatus has a slightly larger angle (60-67°, Fig. lie) than G. bicarinatus {bb-bi° , Fig. lid). Gramma- torcynus has a larger angle than any species of Scom- beromorus excepts, guttatus (60-61°). The ascending process is 33-40% of the total length of the premax- Table 2 Number of teeth in upper and lower jaws of Grammatorcynus. Species Side Min Max X Overall x N Upper jaw G. biUnealiiS L R 14 14 37 36 23.5 24.5 24.0 39 G. bicarinatus L R 14 17 25 24 20.5 20.9 20.7 8 Lower jaw G. bilineatus L R 12 14 32 30 18.6 19.1 18.8 36 G. tiicarinatus L R 16 15 20 23 17.5 17.6 17.5 7 Collette and Gillis. Osteological differences between two species of Grammatorcynus 25 '0mv^ Figure 1 1 Left premaxillae in lateral view, (a.) Scomberomorus lineolatus, Cochin, India, 786mm FL, 2x; (b) Acanthocybium solandri, Miami, FL, 1403mm FL, 1 x ; (c) Grammatorcynus bilineatus, Marshall Is., 424mm FL, 2 x ; (d) G. bicarinatus, Great Barrier Reef, 563mm FL. a Figure 12 Left maxillae in lateral view, (a) Scomber