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Department of Ecology, Evolution and Marine Biology, University of California, Santa Barbara, California 93106 USA; and Department of Ecology and Evolutionary Biology, University of California, Irvine, California 92697 USA
Received for publication March 23, 2000. Accepted for publication August 1, 2000.
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ABSTRACT |
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 TOP ABSTRACT INTRODUCTIONMETHODSRESULTSDISCUSSIONLITERATURE CITED |
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Key Words: Clarkia unguiculata • gender • Onagraceae • pollen : ovule ratio • seed mass • sex allocation • size dependence
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INTRODUCTION |
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TOPABSTRACTINTRODUCTIONMETHODSRESULTSDISCUSSIONLITERATURE CITED |
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), and the pollen : ovule ratio declines over time in the highly selfing Brassica napus (Damgaard and Loeschcke, 1994
). Similarly, the proportion of male relative to bisexual flowers in Anthriscus sylvestris, Leptospermum myrsinoides, and L. continentale (Myrtaceae) increases among upper and outer branch positions (O'Brien, 1994
; Spalik and Woodell, 1994
). In Arisaema triphyllum, younger plants are more likely to be male than female (Policansky, 1981
; Bierzychudek, 1982, 1984
; Clay, 1993
). These results suggest that gender allocation responds to changes in the resource status of individual flowers or whole plants. By contrast, strong ontogenetic changes in gender allocation exhibit no clear temporal pattern in Spergularia marina (Mazer and Delesalle, 1996a, b
), and shifts in gender allocation are seen in some plants independently of variation in resource status (Wolfe, 1992
; Diggle, 1995, 1997
; Ashman and Hitchens, 2000
).
Among conspecific individuals, sex allocation may vary for genetic or environmental reasons. The relative production of ovules and pollen per flower differ among maternal families in the bee-pollinated Campanula rapunculoides (Vogler, Peretz, and Stephenson, 1999
) and among predominantly selfing lines of Senecio vulgaris (Damgaard and Abbott, 1995
), Brassica napus (Damgaard and Loeschcke, 1994
), and Spergularia marina (Mazer and Delesalle, 1996a, b
). In several studies, variation in sex allocation appears to be environmentally induced. For example, allocation to male flowers is positively correlated with an individual's height relative to its conspecific neighbors in Ambrosia artemisiifolia (Lundholm and Aarssen, 1994
; see also Ackerly and Jasienski, 1990
), and the flowers of small plants have higher pollen : ovule ratios than flowers on large plants in Triphyllum erectum and T. grandiflorum (Wright and Barrett, 1999
). In Raphanus sativus, increases in local population density result in the production of relatively male-biased flowers (Mazer, 1992
).
Among populations, variation in sex allocation may result from natural selection, phenotypic plasticity, or genetic drift (Benseler, 1975
; Lovett Doust and Cavers, 1982a, b
; Schoen, 1982
; Delesalle and Mazer, 1995
). Finally, among species, variation in sex allocation is often associated with breeding system and inferred to be the result of adaptive evolution (Queller, 1984
; McKone, 1987
; Ganeshaiah and Uma Shaanker, 1991
).
The recognition that natural selection may discriminate among bisexual genotypes on the basis of their combined male and female fitness motivated the development of theoretical models that predict the conditions that favor particular patterns of sex allocation (Charlesworth and Charlesworth, 1981, 1987
; Charnov, 1982
; Lloyd, 1984
; Lloyd and Bawa, 1984
; Geber and Charnov, 1986
; Charlesworth and Morgan, 1991
; Spalik, 1991
). These models use an evolutionarily stable strategy (ESS) approach to determine the optimal allocation of reproductive resources to male relative to female function; their principles are described in detail elsewhere (Charnov, 1982
; Lloyd, 1984
; Brunet and Charlesworth, 1995
; Klinkhamer, deJong, and Metz, 1997
; Campbell, 1998
). These models share the assumption that, among genotypes with similar amounts of resources, there is an intrinsic negative correlation between the allocation of reproductive resources to male vs. female functions.
One aim of these models is to determine the optimum allocation to male function given specific relationships between allocation and fitness. The optimum is estimated from the shapes of the male and female fitness functions (Fig. 1) and is located where the absolute values of the slopes (or selection gradients; Morgan and Schoen, 1997
) of the two curves are equal. When the shapes of the fitness gain curves depend on plant size or resource status, the optimum male allocation similarly depends on these parameters (de Jong and Klinkhamer, 1989
; Klinkhamer, 1997
). For example, where male reproductive success is a decelerating function of male investment and female reproductive success is a linear (or less rapidly decelerating) function of female investment (the complement of male investment), then as plant size increases the optimum proportional allocation to female function also increases (Fig. 1). This prediction has been supported by the phenotypic correlations observed between plant size and allocation to female function in many monocarpic species (de Jong and Klinkhamer, 1989
; Kudo, 1993
; Klinkhamer et al., 1997).
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) in which allocation to male function is measured as pollen production per flower (Fig. 1). Here, the male and female fitness gain curves measure the number of seeds that are, respectively, sired or produced by a flower with a given allocation to male gametes. As in the whole-plant model, a trade-off between investment in male vs. female function per flower is assumed: controlling for resource status, flowers producing relatively large amounts of pollen must produce relatively few ovules or seeds. This model also assumes that for a given level of pollen production, flowers of high resource status will produce more or higher quality seeds than flowers of low resource status. Given male fitness functions that decelerate more rapidly than female ones, this model predicts that the optimum proportional allocation to female function will be greater in flowers of high than low resource status.
In many species of annual plants, floral resource status may be strongly correlated with floral position on an inflorescence. Annuals often produce flowers over a protracted period along stems that decrease in diameter distally and that bear leaves at basal but not distal nodes. In such species, flowers (and fruits) produced relatively late (and distally) may depend on resources not used first by the fruits and flowers produced earlier. If there is such competition among flowers for resources, then flowers produced later will have access to fewer resources than those produced earlier. This conclusion is consistent with the results of experiments by Solomon (1988)
and Guitián and Navarro (1996)
. If sex allocation among flowers responds to resource status as Brunet and Charleworth's (1995)
model predicts, then successively produced flowers should become relatively male biased. An alternative adaptive explanation may also account for the production of increasingly male-biased flowers over time; protandry in sequentially flowering plants may favor the production of increasingly male-biased flowers over time (Brunet and Charlesworth, 1995
), as observed in Aquilegia caerulea (Brunet, 1996
).
In sum, these models predict that flowers on plants of high resource status should have higher female investment than those on plants of low resource status and that ontogenetic change in floral gender should progress from female- to male-biased flowers. To date, few empirical studies provide data that can be used to evaluate these models. Only two studies of perfect-flowered species have compared gender expression of flowers in small vs. large individuals; both found that small individuals tend to produce male-biased flowers (Damgaard and Loeschke, 1994
; Wright and Barrett, 1999
). Second, all studies to date have examined phenotypic correlations between plant size and sex allocation rather than estimating genetically based correlations. Phenotypic correlations are problematic because they cannot distinguish between environmental and genetic causes of covariation between size and sex allocation (Rose and Charlesworth, 1981
). To detect evidence for the joint evolution (or "coevolution") of resource status and sex allocation, it is necessary to seek evidence for a genetic correlation between the two traits.
The present study examines both ontogenetic variation in sex allocation among flowers within plants and correlations between size and sex allocation among whole plants representing distinct maternal sibships. This allows us: (1) to evaluate whether within a given population sex allocation responds to changes in resource status in a parallel manner within and among plants as predicted by theory, (2) to seek preliminary evidence for a genetic relationship between plant vigor and floral gender expression, and (3) to seek preliminary evidence for genetically based negative relationships between male and female investment per flower and between other floral traits.
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METHODS |
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TOPABSTRACTINTRODUCTIONMETHODSRESULTSDISCUSSIONLITERATURE CITED |
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). The species is highly variable morphologically and with respect to the existence of supernumerary chromosomes and chromosomal rearrangements (Lewis, 1951
; Lewis and Lewis, 1955
; Mooring, 1958, 1960
; Holsinger, 1985
). C. unguiculata is protandrous, herkogamous, and self-compatible but predominantly outcrossing (t 
0.96 in one wild population; Vasek, 1965
). The eight anthers begin to dehisce as soon as the bud opens, exposing pollen that may remain viable for 4 d (Smith-Huerta and Vasek, 1984
). Several days after the bud opens, the stigma becomes exserted and receptive, and petals reach their maximum size. Slopes and canyons of the Santa Ynez mountains (Santa Barbara County, California) support small populations of C. unguiculata. In Rattlesnake Canyon there are several small (N = 10–300 flowering plants) populations of C. unguiculata. The seeds used here were collected from one such population in July 1996 (USGS Santa Barbara Quadrangle, 7.5 min series; Township 4 N; section 3, northeast quadrant). Seeds were collected from 30 maternal plants and stored in envelopes at room temperature.
Experimental design
On 22 January 1997, seeds from each maternal plant were sown in four 18 cm long x 4 cm diameter cylindrical plastic tubes with drainage holes at their base (Stuewe and Sons, Corvallis, Oregon, USA). Tubes were filled to a depth of 10 cm with crushed dry clay; a 7-cm layer of UC xeric mix soil was added and topped with a 0.5-cm layer of gravel. In each tube, 10–20 seeds were sown and watered daily in a UCSB greenhouse. Tubes were placed vertically in plastic racks that rested in a plastic tray filled with 
3 cm of water. As seedlings emerged, they were thinned to one robust seedling per tube. This nonrandom sampling within maternal families should have reduced the potential for maternal environmental effects to influence progeny performance. A total of 116 progeny flowered. Throughout the experiment, plants remained under ambient conditions with 16 h day/8 h night supplemental lighting.
Plants began to flower on 15 April 1997. As each plant flowered, we sampled flowers in sets of three along the plant's primary meristem. The first (most basal) flower in each set was collected when in bud to determine its pollen production before anther dehiscence. The bud was placed in an open 35-mL glass vial in a dust-free cabinet for 48 h. The anthers were then removed, slit longitudinally with a dissecting needle to facilitate dehiscence, and returned to the vial for 1 wk. The vial was then capped until the pollen was counted (see below). The second flower was collected when fully open and frozen in a 2-mL Eppendorf tube. In June–July 1997, the frozen flowers were dissected to determine ovule number per flower, and the area of one petal was estimated as follows. The length of the petal (in millimeters) was measured from its tip to the base of the blade where its width had narrowed to 3 mm; petal width was measured as the maximum width perpendicular to the length; and petal area was estimated in square millimeters as length x width. The third (most distal) flower in each set was hand-pollinated with a mixture of pollen sampled from two plants representing other maternal families. These flowers were allowed to mature to determine the number of viable seeds, mean individual seed mass, and the total seed mass per fruit.
Flowers were collected and pollinated from 15 April to 2 June 1997 and fruits were collected when dry (May–July 1997). For each set of three flowers, the proportion of ovule abortion was estimated as:
The pollen : ovule (P:O) ratio was estimated for each set of flowers as the ratio of the number of pollen grains produced by the anthers in the bud to the number of ovules in the dissected flower. When available, a total of five sets were sampled along the primary branch of each plant. A sixth bud was sampled if present. As each plant began to senesce, water was withheld until the plant was completely dry. The mean dry mass of stem tissue was determined after removing all roots, fruits, and leaves.
Pollen grains were counted using an Elzone 180PC Particle Counter (formerly Particle Data, Inc., currently Micromeritics Instrument Corporation, Norcross, Georgia, USA; Devlin, 1988
). Twenty-four hours before counting pollen, the vial containing the anthers was filled with a known quantity of a 2% NaCl solution (30 mL) and sonicated for 2 min to facilitate the release of pollen from the anthers. Immediately before using the particle counter, samples were sonicated again for 4 min. If the pollen had not fully dehisced from the anthers, they were permitted to soak for another 24 h, by which time all pollen was released from the softened anthers. Using the particle counter, we determined the number of pollen grains in five 0.5-mL samples of this solution, shaking the sample between each count to prevent pollen settling. Based on examination of the solution under a microscope, particles between 75 and 160 µm in diameter were generally considered to represent pollen grains (the exact size range included in a given sample depended on the bounds of the frequency distribution of particles detected by machine). The 0.5-mL samples with the highest and the lowest pollen counts were eliminated from analysis. We estimated the number of grains per flower by multiplying the mean number of grains in the three remaining 0.5-mL samples by twice the number of milliliters of solution in the vial.
Accuracy of the Elzone particle counter
In the course of using the particle counter, we became concerned that it might not distinguish between viable and aborted pollen grains (which we occasionally observed under a microscope). In addition, we discovered that the shaking of the saline solution that was necessary to prevent Clarkia pollen from sinking introduced air bubbles that were counted by the particle counter. We tested the accuracy of the particle counter by comparing the number of pollen grains per flower that it estimated with the number counted by hand. The three-cornered pollen grains of C. unguiculata are large enough (80–160 µm in diameter) that they can be easily counted using a dissecting microscope, although this method takes much more time than the particle counter.
For this comparison, we sampled 38 fresh flowers of C. unguiculata from the same field population used in this experiment. From each flower, we divided the four large and four small anthers into two 2.0-mL Eppendorf tubes, each containing two large and two small anthers. The tubes were left uncapped for 1 wk to facilitate anther dehiscence.
The tubes with the pollen to be hand-counted were treated as follows. One milliliter of 50:50 ethanol : glycerin solution was added and the contents vortexed thoroughly. From 19 of these samples, 20 µL were added to each of five glass slides along with one drop of Alexander's stain (Alexander, 1980
). On the slide surface, the solution was mix-pipetted with a clean micropipette tip and then covered with a cover slip. The slides were etched with a 16 x 16 mm grid to facilitate counting. On each slide, all "normal" and "aborted" grains were counted. Grains assumed to be normal were those with turgid walls, of full size, and that had absorbed the red stain. Grains assumed to be aborted were deflated, a maximum of 80 µm in diameter, and remained green and translucent. The number of grains produced by an entire flower was estimated by calculating the mean number of grains per slide and multiplying this value by 100. For these 19 samples, we estimated pollen production per flower using sample sizes of both three and five slides per sample. The correlation between the mean number of pollen grains per 20 µL solution using three vs. five slides was very high (r = 0.96, N = 19, P < 0.0001). From the remaining 19 pollen samples we limited the sample size to three slides. The pollen in each of the Eppendorf tubes containing the other four anthers was transferred to a 35-mL glass vial and then counted using the particle counter as described above.
Among the 38 pollen samples, there was no significant difference among methods in the estimated number of pollen grains per flower (F2,11 = 2.66, P > 0.07). The mean number of normal grains per flower estimated by hand was 12 005 (SD = 3672, range: 4533–18 733); the mean number of all grains (normal and aborted) per flower estimated by hand was 12 996 (SD = 3674, range: 6000–19 993); and the mean number estimated by the pollen counter was 14 189 (SD = 4935, range: 5101–27 229). The mean proportion of aborted grains per sample was 0.08 (SD = 0.09). Assuming that the hand-counted estimates of pollen production are more accurate (they contained no bubbles), the particle counter overestimated the total number of grains per flower by 9% and the number of normal grains by 18%.
To determine whether the estimates derived from the particle counter accurately predicted the values estimated by hand, a regression was conducted. The regression of the number of normal grains per flower estimated by hand on the number estimated by the particle counter was highly significant (y = 0.32x + 7719, df = 37, P < 0.005), but the correlation coefficient (r) was only 0.45 (though significant; P < 0.005). The relatively low value of r indicates that the estimates of pollen production per flower and the pollen : ovule ratios reported below are not highly accurate. Although the particle counter introduced error, it is unlikely to have introduced bias into our data set. The comparisons of the mean pollen production per flower and mean pollen : ovule ratios among the sets of flowers and maternal families sampled in this experiment should not be compromised even though the values of these variables were overestimated by the particle counter.
Statistical analysis and interpretation
Variation among maternal families in floral traits—Phenotypic differences among the progeny of different field-collected maternal plants may be due to additive or nonadditive genetic variation and/or due to environmentally induced maternal effects. Nonadditive sources of genetic variation among families include dominance, epistasis, or differences among families in inbreeding depression. Plants in at least one wild population of C. unguiculata exhibit outcrossing rates close to 100% (Vasek, 1965
), however, suggesting that interfamily variation in inbreeding depression is likely to be very low.
Raising the progeny under uniform greenhouse conditions and free of competition increases the likelihood that differences observed among the progeny of distinct maternal families have a genetic basis, but additive and nonadditive sources of variation cannot be distinguished. Also, maternal environmental effects on progeny phenotype can create the false appearance of heritable variation among families. In all studies based on maternal families, inferences concerning the genetic basis of differences among maternal sibships must be viewed with caution. By using only robust seedlings from each family, however, we aimed to minimize environmentally induced maternal effects. The sixfold variation observed among families in final vegetative size (see below) suggests that maternal families exhibited significant genetic variation in resource acquisition or in resource use efficiency, at least under greenhouse conditions.
To evaluate whether maternal families differed in the mean values of their floral traits, we conducted one-way ANOVAs (Statview, SAS Institute, 1999
) using data from each floral position separately. Pollen per flower, mean individual seed mass, and petal area were lognormally transformed to improve normality.
Family-based covariation among floral traits
Genetic covariation among maternal families was estimated using maternal family means of floral traits, where each family's mean was calculated from the phenotypic means of the individuals representing it. Although we aimed to collect five sets of flowers from the primary flowering stem of all individuals, not all stems produced five buds, flowers, and fruits. Many individuals contributed no more than four sets. In the case of pollen production, many of the first buds sampled were eliminated from the analysis because the pollen did not fully dehisce from their anthers (from the first buds, the anthers had not been slit prior to drying). Therefore, to control for phenotypic variation among individuals and families that might be due to the combination of unequal flower production (across all positions) and ontogenetic effects on floral traits, we first created a balanced data set with which to calculate maternal family means. Each record in this data set represented an individual plant, and for all traits except for pollen production, the data set included only those individuals that contributed the first four sets of flowers and fruits (no other sets were included). In the case of pollen production, many buds in the first set were eliminated from the analysis because the pollen did not fully dehisce from their anthers. Consequently, to estimate family means for pollen production per flower and the pollen : ovule ratio, we analyzed data from all plants contributing buds from the second, third, and fourth sets (no other sets were included).
When estimating covariances, environmentally induced differences among maternal families in plant vigor could bias among-family correlations towards positive values. We therefore controlled for differences among families in plant size by estimating partial correlation coefficients between floral traits, controlling for stem biomass. The balanced data set and the partial correlations allowed the detection of among-family correlations between floral traits independent of stem biomass (SAS, 1989
) and floral position. All traits except for the P:O ratio and the proportion of ovule abortion were lognormally transformed to reduce heteroscedasticity.
These analyses were used to seek evidence for genetic covariation between components of reproduction, and, in particular, to seek evidence for trade-offs between male and female investment. In accordance with the view that corolla size contributes more to male than to female fitness in animal-pollinated outcrossing taxa (Charnov, 1979
; Willson, 1979
; Bell, 1985
), we considered pollen production and petal area to represent male investment. The fact that C. unguiculata petals do not reach maximum size until the anthers have dehisced (i.e., when the flower is just beginning its female phase) may challenge the view that petal size contributes more to male than to female fitness in this species. At the whole plant level, however, if large-flowered genotypes attract more pollen vectors than small-flowered plants, then large-petalled flowers may contribute disproportionately to male fitness.
These family means were also used to calculate residuals of ovules per flower, pollen per flower, seeds per flower, and petal area on stem biomass. Regressions between pairs of these residuals were conducted as a second means of detecting trade-offs between male and female reproductive components independent of plant size.
Ontogenetic change in floral traits
Ontogenetic change in sex allocation was observed by examining how traits related to female and to male function changed as one progresses from early to late flowers along the primary flowering stem. Temporal change in floral traits was detected statistically using a repeated measures ANOVA (SAS Institute, 1989
). Maternal family and individual were independent variables, with floral traits measured repeatedly within individuals.
Size-dependent change in floral traits among maternal families and among individuals
To detect size-dependent changes in per-flower sex allocation, we examined the regressions among maternal family means between the mean floral phenotype of each trait and mean vegetative stem biomass (Statview, SAS Institute, 1999
). To control for phenotypic differences among maternal families that might be due to the combination of unequal flower production per family and ontogenetic change in floral traits, we used constructed two balanced data sets.
The first data set consisted of maternal family means calculated using only those individuals for which data from the first four sets of flowers and fruits were available; family means for pollen production and for the P:O ratio were based on individuals contributing the second through fourth sets of buds and flowers. This data set included information from 85 individuals for all traits except pollen, for which 63 individuals were included.
The second data set consisted of maternal family means calculated using data from all individuals, but only from the first two sets of flowers and fruits; family means for pollen production were based on buds from the second and third sets. This data set included information from 114 individuals for each trait. The first data set has the advantage of including more flowers per individual, while the second data set includes more individuals per family. These data sets allowed us to determine whether, controlling for flower number and position, there is a relationship between sex allocation and vegetative size among maternal families. If variation among maternal families in size is due to genetic variation, a significant regression coefficient indicates a genetic basis to the relationship between size and phenotype.
Regressions were also conducted using individual plant means (including individuals contributing the first four sets of flowers only) to determine whether phenotypic regressions, in which size differed among individuals by a factor of 23, differed from the among-family regressions, where size differed among family means by a factor of six.
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RESULTS |
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TOPABSTRACTINTRODUCTIONMETHODSRESULTSDISCUSSIONLITERATURE CITED |
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4800 to 29 000 pollen grains (mean = 12 406, SD = 4075, N = 354 buds). The pollen : ovule ratio ranged from 56 to 427 (mean = 146, SD = 55.08, N = 351 buds and flowers). Fruits produced 5–95 viable seeds per fruit (mean = 47.4, SD = 21.96, N = 451 fruits), with mean individual seed mass per fruit ranging from 0.18 to 1.10 mg (mean = 3.97, SD = 1.33, N = 439 fruits). The area of single petals ranged from 63 to 195 mm2 (mean = 121.0, SD = 21.03, N = 491 flowers). Among all individuals monitored in this experiment that contributed data from both the first and second sets of flowers sampled (or the second and third bud, for pollen), mean ovule production per flower ranged from 57 to 118 (N = 116 individuals), and mean pollen production per flower ranged from 6585 to 25 237 grains (N = 70). The pollen : ovule ratio among individuals contributing data from the second and third sets of flowers and buds varied from 80 to 346 (N = 58). Individuals produced a mean of 5.5–86.5 viable seeds per fruit (N = 114) with mean individual seed mass ranging from 0.19 to 0.85 mg (N = 113). The mean area of a single petal varied among individuals from 85.3 to 181.2 mm2 (N = 116). Stem biomass among individuals ranged from 0.10 to 2.31 g (N = 116).
Variation among maternal families in floral traits
For most traits and floral positions, only female traits differed significantly among maternal families (Table 1). For at least three of the five floral positions, families differed significantly with respect to the mean number of ovules per flower, the number of seeds per fruit, total seed mass per fruit, and the estimated proportion of ovule abortion. Significant differences among maternal families were detected for mean individual seed mass only for the first and second fruits sampled. Maternal families differed with respect to the pollen : ovule ratio only for the last flower sampled.
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Ontogenetic change in floral traits
Repeated-measures ANOVA detected significant effects of floral position on ovule number per flower, the P:O ratio, viable seeds per fruit, total seed mass per fruit, and mean individual seed mass (Table 3). Flowers produced relatively late produced significantly fewer ovules than those produced earlier (Fig. 2). Similarly, seed number and total seed mass per fruit declined significantly from basal to distal positions. The reduction in seed number in distal positions was so great that, even though total seed mass per fruit declined distally, mean individual seed mass increased towards the distal end of the flowering branch.
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Size-dependent change in floral traits among maternal families and among individuals
Stem mass provided a reliable measure of growth, reproduction, and apparent plant vigor. Stem mass was positively correlated with lifetime flower production, branch number, and total branch length, and negatively correlated with branch length per flower (Fig. 3). Families of relatively large individuals produced more flowers over their lifetime and more flowers per unit branch length than families of small individuals. Among families, lifetime flower production increased proportionately with plant size (the slope of the ln–ln regression of lifetime flower production on stem biomass is not significantly different from 1.00).
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DISCUSSION |
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TOPABSTRACTINTRODUCTIONMETHODSRESULTSDISCUSSIONLITERATURE CITED |
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Differences between previous studies and the current study
This study differs from previous ones in several ways. First, we examined size-dependent sex allocation at two levels: (1) as plants grew and (2) among plants and maternal families differing greatly in whole-plant size. Other studies to date of floral sex allocation have been conducted either within or among plants. This approach allowed us simultaneously to test the two predictions just mentioned and to ask whether within a given population floral traits are equally sensitive to ontogenetic factors and to whole-plant attributes. We found that ontogenetic variation in sex allocation was much greater than variation among plants of different size. The lack of variation detected among individuals and families in sex allocation is consistent with the growing view that natural plant populations harbor greater genetically based variation in resource-garnering ability among individuals than in the proportional allocation of resources to male vs. female investment (Campbell, 1997
; Fenster and Carr, 1997
).
Second, this study was unusual in its focus on gender allocation within bisexual flowers. Most previous studies have been conducted either on monoecious (or andromonoecious) species, where the ratio of male to female or male to hermaphroditic flowers is used to estimate sex allocation, or on perfect-flowered species, where the ratio of lifetime flower production (a proxy for male investment) to lifetime fruit or seed production (female investment) is used to estimate sex allocation. Within-flower changes in gender allocation are distinct from those measured at the whole-plant level in that the former require developmental "decisions" to occur as each flower is produced. By contrast, the change from female to male flower production in monoecious species occurs once during the development of an inflorescence or individual, and changes in the fruit to flower ratio depend on post-fertilization abortion occurring throughout fruit development. While numerous studies of ontogenetic changes in reproductive components of hermaphroditic flowers have been conducted (Ashman, 1992
; Stoecklin and Favre, 1994
; Vogler, Peretz, and Stephenson, 1999
; Medrano, Guitian, and Guitian, 2000
), detailed studies of both male and female primary sexual traits are relatively rare (Mazer and Delesalle, 1996a, b
; Lehtila and Strauss, 1999
; Ashman and Hitchens, 2000
).
Third, by raising offspring derived from maternal families, we sought preliminary evidence of genetic variation in both floral gender and plant vigor. One advantage of examining size-dependent sex allocation among maternal families raised under uniform conditions is that an observed correlation between plant size and gender is likely to have a genetic basis. By contrast, phenotypic associations between plant size and sex allocation observed in a heterogeneous field environment (e.g., Wright and Barrett, 1999
) are more likely to be environmentally induced. Establishing a genetic basis for the association between size and gender is important because only such genetic associations can be the outcome of natural selection (and not simply environmentally induced). Here we detected no evidence for a genetic basis to size-dependent sex expression.
The lack of variation in floral sex allocation among plants that differ greatly in biomass suggests that resources alone might not be the most important cause of position effects on floral sex allocation within plants. Architectural or developmental changes associated with floral position may have differential mechanical effects on stamen vs. carpel development, causing position effects on sex allocation independent of resource availability. In addition, differential movement or production of hormones throughout an inflorescence or flowering branch could affect the relative development of male vs. female floral parts. Moreover, differences among flowers in gene expression might be position dependent, resulting in phenotypic changes along the axis of an inflorescence. While some experimental evidence indicates that resource availability does vary with floral position (Solomon, 1988
; Guitián and Navarro, 1996
), developmental responses to position effects on vasculature, hormone availability, or gene expression cannot be ruled out as proximal causes for position-dependent sex allocation. Indeed, the expression of some genes have been found to be associated with flower or inflorescence position (Mazzucato et al., 1999
; Yu et al., 1999
), potentially accounting for observed position effects on sex allocation.
Implications for models of size-dependent sex allocation
If the observed changes in sex allocation per flower within plants and among families had been quantitatively similar and were the result of adaptive evolution, this would suggest that the shapes of the fitness gain curves for high- vs. low-resource flowers mirror those for high- vs. low-resource individuals in C. unguiculata. In this study, however, ontogenetic change in floral sex allocation was much greater than size-related change among families.
If the changes in floral sex allocation observed here are adaptive, this would suggest that the male and female fitness gain curves for high- vs. low-resource status flowers within C. unguiculata plants differ qualitatively from those for flowers produced by high- vs. low-resource status individuals. Our results suggest that as one progresses from basal to distal flowers within plants, the optimum proportional investment in male function increases, as predicted by Fig. 1. In contrast, the absence of size-dependent P:O ratios observed among families suggests that the per-flower male and female fitness functions for small vs. large plants do not generate size-dependent optima for primary sex allocation. In this study we did not estimate sex allocation at the level of whole plants, so it is unknown whether the fruit : flower ratio increases (becomes more female-biased) with increasing size (see Aker, 1982
; Watkinson, 1982
; de Jong and Klinkhamer, 1989
; Klinkhamer and de Jong, 1993
).
Previous evidence for resource-dependent sex allocation
Previous studies of size-dependent patterns of sex allocation in monocarpic, perfect-flowered plants have suggested that the shapes of male and female fitness functions may be resource dependent. For example, Klinkhamer and de Jong (1987, 1993)
found strong correlations between plant size and components of male and female reproduction in Cynoglossum officinale (Boraginaceae). In the field, small plants have higher rates of ovule abortion and fewer seeds per fruit per gram of dry mass than large plants. In addition, small plants produce more flowers per gram of dry mass than large plants. Both patterns indicate increased proportional male investment among small plants.
Similar size-dependent changes in whole-plant sex allocation have been found in the perennial Yucca whipplei (Agavaceae) (Aker, 1982
). Relative to the size of the basal rosette, lifetime flower production was higher in small plants than in large ones. In addition, mean seed mass increased with plant size in one of two populations studied. Vulpia fasciculata (Watkinson, 1982
) exhibited a similar pattern: lifetime seed production increased more rapidly with plant size than did total flower production. In a review of 34 studies representing 31 hermaphroditic, monocarpic species, de Jong and Klinkhamer (1989)
report that in 28 cases there was a positive phenotypic correlation between size and some standardized measure of female investment (sensu Lloyd and Bawa, 1984
). These results are consistent with the view that the shapes of male and female fitness curves are similar to those shown in Fig. 1.
Male vs. female investment: no negative genetic correlations
We detected no evidence for a negative correlation among family means between male and female components of reproduction, even when controlling for total plant size (Table 2). The absence of significant negative correlations between male and female investment suggests that genetic variance in the resource-garnering ability (or resource use efficiency [RUE]) of whole plants may be high enough to obscure a negative genetic covariance between resources invested in male and female function (van Noordwijk and de Jong, 1986
; Houle, 1991
). Indeed, when the genetic coefficients of variation (CVg) based on maternal family means of aboveground stem biomass and the pollen : ovule ratio are estimated (Houle, 1992
), it appears that there is higher genetic variation in resource acquisition (or RUE) than in gender expression. The CVg for ln-transformed stem biomass was 36.6% while that of the ln-transformed pollen : ovule ratio was 1.5%. This result supports the view that the relatively high variation among families in their levels of resource acquisition or RUE makes it difficult to detect negative relationships between components of reproduction. However, the inability to detect a trade-off between male and female investment even when stem biomass is controlled statistically suggests either that high variation among maternal families in resource status cannot alone explain the lack of a negative correlation or that stem biomass is not a good assay for the resources used for gamete production.
The one trade-off between male and female gender expression observed here is the negative correlation between the P:O ratio and seeds per fruit (r = -0.32, P > 0.14; Table 2), which merits further exploration. Significant negative relationships were detected between some components of female reproduction (e.g., seed number per fruit vs. mean individual seed mass), suggesting that resources not allocated to one trait may be diverted to another. In the current study, however, this kind of diversion did not generally occur between male and female reproductive components.
Several other studies have similarly detected no negative genetic correlations between male and female reproductive components in wild species with bisexual flowers. Mossop, Macnair, and Robertson (1994)
cultivated ten clones of Mimulus guttatus from a natural population, providing measures of anther mass (which is correlated with pollen production), pollen viability, corolla mass, ovary size, and fruit mass. These traits were measured under two treatments: one in which all flowers were pollinated (high-resource demand) and another in which flowers were removed after anthesis (low-resource demand). In spite of high between-clone variation, there was no evidence for a trade-off between anther mass and ovary size, even when corolla size was controlled for statistically. Similarly, plants producing larger quantities of viable pollen grains per flower exhibited no deficits in flower production, ovary or fruit size, nor did they show an increased rate of decline in the phenotypic value of these traits as plants age. In another study of M. guttatus (Robertson, Diaz, and Macnair, 1994
), quantitative genetic parameters were estimated for 20 floral traits. Although both pollen quality and ovule production exhibited significant levels of additive genetic variation, there was no detectable genetic correlation between them. A positive correlation between pollen and seed production, however, was detected, suggesting that variation among clones in resource-garnering ability may exceed variation in the proportional allocation of resources to male vs. female investment. In a third study of M. guttatus, Fenster and Carr (1997)
raised full-sib families representing a highly selfing and an outcrossing population and measured ovule and pollen production per flower, pollen size, and corolla size. No negative genetic correlations were detected among these traits, even when corolla size was controlled statistically. The authors suggested that the positive correlations observed between components of male and female reproduction may be due to variation among genotypes in resource-garnering ability.
To test the assumption that increased allocation to male function results in reduced female function in Ipomopsis aggregata, Campbell (1997)
examined floral traits among 32 paternal half-sib families and 229 maternal half-sib families. She detected no significant negative genetic correlation between stamen biomass and any measure of male or female investment: corolla mass, calyx mass, pistil mass, and total seed mass per fruit. Indeed, among paternal family means, stamen biomass was positively correlated with corolla, pistil, and seed mass. These results again suggest that genetic variation in resource-garnering ability may obscure trade-offs between the allocation of resources to male vs. female function. A similar conclusion was drawn in a study of tristylous Lythrum salicaria (O'Neil and Schmitt, 1993
).
By contrast, several studies have detected evidence for genetically based trade-offs between male and female investment at the level of individual flowers. Artificial selection over three generations to reduce anther number per flower resulted in a compensatory increase in ovule production per flower in Spergularia marina, and selection to increase ovule number per flower resulted in a decline in anther number (Mazer, Delesalle, and Neal, 1999
). A study of Campanula rapunculoides detected a significant negative correlation among 11 clones between ovule and pollen production per flower (Vogler, Peretz, and Stephenson, 1999
). Among hermaphroditic genotypes in the gynodioecious Fragaria virginiana, pollen production per flower was negatively correlated with the probability that a flower would develop into a fruit (Ashman, 1999
). Rameau and Gouyon (1991)
reported negative genetic correlations between seed mass and the number of viable pollen grains produced in horticultural clones and hybrids of Gladiolus. Finally, Atlan et al. (1992)
reported a negative correlation among maternal family means between the number of germinating seeds per fruit and the number of full pollen grains per flower in Thymus vulgaris.
Identifying the adaptive significance of size-dependent sex allocation
It is not possible to assert whether observed changes in sex allocation are adaptive without knowing the shapes of male and female fitness functions depicted in Fig. 1. Predicting the optimal allocation to male (or female) function requires precise estimates of the curvature of these functions, which may be impossible to obtain without huge sample sizes and a large number of molecular markers. Moreover, the shapes of male and female fitness gain curves may depend on the quality of the environment in which plants grow: a possibility that has not yet been explored. It may therefore be unrealistic to expect to estimate these parameters for a wide range of species under field conditions.
An alternative approach to detecting the adaptive significance of size-dependent sex allocation would be to compare patterns of gender change in species with different mating systems or pollen vectors. For example, wind-pollinated species may be more likely to exhibit increases in proportional allocation to male function as plant size increases than are entomophilous species (Bickel and Freeman, 1993
; Dajoz and Sandmeier, 1997
). This has been proposed to be the result of the increased dispersal ability of pollen with increased plant height in wind-pollinated taxa. One would also predict that completely autogamous taxa should not exhibit size-dependent sex allocation, as all flowers contribute equally to both male and female function. By such comparisons, it may be easier to find support for the argument that size-dependent sex allocation is the adaptive outcome of natural selection than to rely on the estimation of fitness functions.
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FOOTNOTES |
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2 Author for correspondence (mazer{at}lifesci.ucsb.edu
).
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LITERATURE CITED |
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TOPABSTRACTINTRODUCTIONMETHODSRESULTSDISCUSSIONLITERATURE CITED |
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