The strictly allopatric model of speciation makes definable predictions on the pattern of divergence, one of which is the uniformity in the divergence time across genomic regions. Using 345 coding and 143 intergenic sequences from the African great apes, we were able to reject the null hypothesis that the divergence time in the coding sequences (CDSs) and intergenic sequences (IGSs) is the same between human and chimpanzee. The conclusion is further supported by the analysis of whole-genome sequences between these species. The difference suggests a prolonged period of genetic exchange during the formation of these two species. Because the analysis should be generally applicable, collecting DNA sequence data from many genomic regions between closely related species should help to settle the debate over the prevalence of the allopatric mode of speciation.
THE allopatric mode of speciation is the tenet of the neo-Darwinian view of speciation (Mayr 1963). In this view, a geographical barrier preventing gene flow is a prerequisite for speciation. Without such barriers, gene exchanges during the process of species formation would obstruct the process as such exchanges would destroy the adaptive gene complexes and obliterate the accumulated differences between nascent species. On the other hand, there is no compelling population genetic reason why divergent adaptation cannot proceed in the presence of continuous gene flow (e.g., Navarro and Barton 2003). A most common mode may be parapatric speciation when nascent species are geographically connected by gene flow (Mayr 1963; Endler 1977). The extreme form of gene flow is represented by sympatric speciation (Dieckmann and Doebeli 1999; Kondrashov and Kondrashov 1999). Parapatric speciation may best be envisioned at the genic level (Wu and Ting 2004) where portions of the genome progressively become divergently adapted and hence nonexchangeable between nascent species. The genealogical history of the genome would therefore be mosaic with disparate divergence time among different loci. While most previous tests of the allopatric vis-à-vis parapatric mode of speciation were based on ecological or biogeographical considerations (Endler 1977; Butlin 1998; Coyne and Price 2000), only a few studies have utilized multiple DNA sequences (Kliman et al. 2000; Machado et al. 2002) for this purpose. This analysis represents a genome-wide perspective on the same issue.
In strict allopatry, all the genes in the genome should have the same divergence history (t in Figure 1A) but vary in the coalescence time, which is exponentially distributed with mean equal to 2Ne (Ne being the effective population size at the time of speciation, Figure 1A). We discuss more complex forms of allopatric speciation later. Note that time is measured in units of generation throughout this report. A large variance in DNA divergence can be due to either variation in t across loci or a larger-than-estimated Ne, both of which can enhance the variance in divergence among loci. Although there are many studies for estimating Ne, they all assume constant t across loci, or strict allopatry, precisely what we wish to test. Interestingly, when t is assumed to be a constant, the estimated Ne's for the ancient species are usually far larger than those for the extant populations (Ruvolo 1997; Takahata and Satta 1997; Chen and Li 2001; Yang 2002; Wall 2003). These studies thus hint at the possibility of nonconstant t.
In this study, we compare coding and intergenic regions for their evolutionary dynamics during speciation. In allopatry, these two types should have the same dynamics but, under the parapatric model of speciation, could have very different histories. Figure 1B illustrates this point, on the assumption that coding sequences are more likely than intergenic sequences to be associated with hybrid incompatibility or differential adaptation. The potential for coding regions to successfully move across nascent species boundaries may be curtailed early (Figure 1B). On the other hand, intergenic regions, experiencing less impediment to their trafficking between nascent species, should continue to be exchangeable until openings in the reproductive barrier are completely sealed. This contrast has been reported for Drosophila between DNA sequences at or near a speciation gene (Ting et al. 2000). A recent report also assumes that the common ancestors of human and chimpanzee went through a period of parapatry (Navarro and Barton 2003). However, the observations were reanalyzed in light of outgroup data and were suggested to result from events unrelated to speciation (Lu et al. 2003; Navarro et al. 2003).
MATERIALS AND METHODS
We collected 98 common chimpanzee (Pan troglodytes) sequences from the GenBank database, 93 from the 5′-conseusus sequences of Sakate et al. (2003), 19 newly determined full-length cDNA sequences from Ryuichi Sakate and Momoki Hirai (University of Tokyo), and 135 genomic sequences of chimpanzee chromosome 22 corresponding to Ensembl genes of human chromosome 21. Seventy-six gene sequences of gorilla were collected from GenBank. We removed from our analysis MHC sequences, whose genealogy was deeper than human-chimpanzee divergence due to strong balancing selection (Satta et al. 1999). DDBJ/EMBL/GenBank accession numbers of newly determined sequences are AB188273, AB188274, AB188275, AB188276, AB188277, AB188278, AB188279, AB188280, AB188281, AB188282, AB188283, AB188284, AB188285, AB188286, AB188287, AB188288.
The sequences were aligned by using the ClustalW program (Thompson et al. 1994) and corrected by visual inspection. Numbers of synonymous and nonsynonymous substitutions were estimated by the method of Li (1993) with equal weighting among pathways for multiple substitutions in a codon (Pamilo and Bianchi 1993). Fifty-three intergenic sequences were obtained from Chen and Li (2001). Ninety pairs of 2-kb intergenic sequences of human and chimpanzee, which are at least 10 kb apart from genic regions annotated by Ensembl, were obtained from genomic sequences of human chromosome 21 and chimpanzee chromosome 22. Numbers of substitutions of intergenic regions were estimated by using Kimura's two-parameter method (Kimura 1980).
Maximum-likelihood estimate of divergence time and ancestral population size:
We designate ki as the number of synonymous changes for the ith sequence [either coding sequence (CDS) or intergenic sequence (IGS)]. The probability of observing ki is given by 1where τi = 2tui and θi = 4Neui (Equation 5 in Takahata and Satta 1997). li is the length of sequence i and ui is the per-nucleotide substitution rate for the ith sequence. Equation 1 has two components—the Poisson distribution in the divergence portion (τi = 2tui) and the “mismatch distribution” in the coalescence portion (θi = 4Neui), where the absence of intragenic recombination is assumed.
Without the outgroup:
We first assume that the substitution rate for CDS (and separately for IGS) is uniform across m loci when the outgroup sequences are not available for calibrating the variation in the mutation rate. Let τ = 2tu and θ = 4Neu. The log-likelihood for Equation 1 becomes
The maximum-likelihood estimates (MLEs) of τ and θ were found by numerical iteration.
With the outgroup:
We now use sequences from an outgroup species, say the orangutan, to filter out the variation in ui. Let the divergence time between human and the outgroup be T and the substitution number of the ith locus between these two species be Ki. We assume that Ki = 2liTui without considering the coalesence component because, if T is sufficiently large, the impact of 2Ne should be insignificant. Noting that the divergence time between human and chimpanzee is t, we can now replace τi and θi with (t/T)(Ki/li) = α(Ki/li) and (2Ne/T)(Ki/li) = β(Ki/li), respectively, in Equation 1. α = t/T and β = 2Ne/T are the two parameters to be estimated by MLE.
The log-likelihood function is and the MLE of the two parameters, α and β, can be found by numerical iteration.
In the allopatric model, the divergence of all genes is fixed at t (Figure 1A), while in the parapatric model (Figure 1B), the divergence time is uniformly distributed between t and 2t. For 1000 loci, random integers in the range of 500–1000 are generated for the number of sites. Coalescent times are generated as exponential distribution whose mean is 2Ne while Ne satisfies γ = t/2Ne = 10. t corresponds to 1% nucleotide divergence with 1.5 × 10−8 substitutions per generation per site. The number of substitutions is assigned according to the Poisson distribution where the mutation rate is uniform among loci. A confidence interval of 95% was calculated by 1000 iterations.
Congruence between gene genealogy and species phylogeny:
Seventy-six gene sequences of human, chimpanzee, gorilla, and orangutan were used for the congruence test. The observed genealogies can be (M, M, m), (m, M, M), or (M, m, M) (see Figure 1C), where m and M are the ancestral and derived variants, respectively. (m, M, M) and (M, m, M), where gorilla shares the derived variant with either chimpanzee or human, are incongruent with the species phylogeny, in which human and chimpanzee are the closest relatives. In our analysis, we first treated each site independently. All CG to TG and CG to CA substitutions were masked from the analysis because of the very high rate of changes at such CpG sites, which often results in genealogies incongruent with the species phylogeny. Adjacent variant sites within the same locus that show the same genealogical pattern are counted as one segment. A locus may have more than one segment showing different phylogenetic patterns, presumably due to recombination.
We designate the observed number of segments that show the pattern of (M, M, m), (m, M, M), and (M, m, M) a1, b1, and c1, respectively, for IGS. Likewise, the numbers are a2, b2, and c2, respectively, for CDS. Under the null hypothesis that t′/2Ne′ is the same between coding and intergenic regions (Figure 1C), the likelihood ratio R is where a3 = a1 + a2, b3 = b1 + b2, c3 = c1 + c2, n1 = a1 + b1 + c1, n2 = a2 + b2 + c2, and n3 = a3 + b3 + c3 (derived from Equation 7 in Wu 1991).
RESULTS AND DISCUSSION
We estimate τ = 2tu and θ = 4Neu, where u is the per-nucleotide substitution rate, by the maximal-likelihood (ML) method (Takahata and Satta 1997; see materials and methods). Our objective is to test if t is the same between coding and intergenic regions. However, because u many not be the same between two regions, we define γ = τ/θ = t/2Ne and test if γ is the same between the two regions. γ is the relative divergence accrued after, vis-à-vis before, speciation and should be constant under the null hypothesis of allopatry.
To know how parapatry might affect the estimation of γ when allopatry (i.e., constant t across loci) is incorrectly assumed, we carried out computer simulations. In the simplest case, the divergence time in the allopatric model is fixed at t (Figure 1A), while the divergence time in the parapatric model is uniformly distributed between t and 2t (Figure 1B). More complex simulations have been done but the results can be qualitatively stated as such: parapatry generally results in the underestimation of γ (= t/2Ne). Even when the true γ is 50% larger in parapatry than in allopatry, as in the case of Figure 1, the estimated numbers are nevertheless in the opposite direction (Table 1). The reason for this seemingly paradoxical result is that, under parapatry, the estimate of 2Ne is greatly inflated to account for the variation in the level of divergence among loci. Hence, we expect γ to be underestimated when allopatry is incorrectly imposed on data that have a variable divergence time. Coding sequences probably fit this characterization better than intergenic sequences (Figure 1B).
We used 345 CDSs and 143 IGSs from human and chimpanzee and conducted the likelihood-ratio test between the two hypotheses, γCDS = γIGS = γ0 and γCDS ≠ γIGS, where γCDS and γIGS are MLEs for the CDS and IGS, respectively (Takahata and Satta 1997). Under the null hypothesis, the MLE of γ0 is 1.89 and the log-likelihood value is −1098.588 (Table 2). Under the alternative hypothesis, the MLEs for the two regions are γCDS = 1.31 and γIGS = 2.45 and the log-likelihood value is −1096.226 (Table 2). The likelihood-ratio test between the two models yields a significant result (P = 0.027). Because the variation among loci in the number of CpG sites, which exhibit high mutability (Hellmann et al. 2003), may have an impact on our estimation, we reestimated γ by masking all CG to TG and CG to CA substitutions. The likelihood-ratio test leads to the same conclusion (P = 0.006, see supplementary Table 1 at http://www.genetics.org/supplemental/). Strictly speaking, because Ne may be smaller for the coding than for the intergenic region, as the former is generally less variable than the latter (Pluzhnikov et al. 2002), the null hypothesis should be γIGS ≤ γCDS, making our test conservative. The null hypothesis of γIGS ≤ γCDS is thus rejected.
For the method to be of general use in testing allopatric speciation, the need for DNA sequences should not exceed what we used above. A need for >500 sequences would make the method impractical for most specie pairs. Nevertheless, between human and chimpanzee, 7645 orthologous sequences are available (Clark et al. 2003) to back up the above analysis. For this large dataset, γCDS is 1.20, which leads to an even more significant likelihood ratio (P = 0.0003, see supplementary Table 1). Above 500 sequences, an increase in sample size >500 in this case appears to yield a diminishing return.
To standardize the divergence measure and make it independent of the underlying mutation rate, we also calibrate the human-chimpanzee divergence against the divergence between these two species and an outgroup. We were able to use only 76 CDSs and 53 IGSs from human, chimpanzee, and orangutan for this purpose. It is assumed that the level of divergence between human and orangutan is a function of their divergence time, T, without much influence by the ancestral polymorphism, the contribution of which should be relatively small here. The key parameters are now α = t/T and β = 2Ne/T (see Figure 1 and materials and methods). By doing so, γ (= α/β) was estimated to be 1.55 and 37.3 for the CDSs and IGSs, respectively. While the estimates are different from those of Table 2 due to both the small sample sizes and the inherent variability in the estimation of γ (see Table 1), the general trend of γIGS ≫ γCDS is observed.
When calibrated against the divergence from the orangutan, the divergence in CDS and IGS between human and chimpanzee can in fact be directly compared since the governing parameters, α = t/T and β = 2Ne/T, depend only on the common elements, t, T, and 2Ne. For each locus, we therefore compute the relative divergence dR = dhc/[(dho + dco)/2], where dhc, dho, and dco are the levels of divergence between human and chimpanzee, human and orangutan, and chimpanzee and orangutan, respectively. The mean of dR is 0.522 for CDS and 0.404 for IGS (P = 0.030) while the variance of dR is 0.166 for CDS and 0.037 for IGS (P < 10−7). The results suggest that, on average, coding regions have deeper genealogy than intergenic regions and the variation is larger in the former than in the latter, as hypothesized in Figure 1.
The analysis of Table 2 has also been applied to the divergence between gorilla and either human or chimpanzee (node b of Figure 1C). By using 76 coding and 53 intergenic sequences the null hypothesis of allopatry cannot be rejected (P = 0.950 for human-gorilla and P = 0.224 for chimpanzee-gorilla). Although the results are not significant, the chimpanzee-gorilla comparison appears to be very different from the human-gorilla divergence. In the former, γIGS > γCDS and the difference is larger than that in the human-chimpanzee comparison (Table 2). Given the small number of sequences from gorilla, there is little statistical power to resolve the issue at this moment. Nevertheless, chimpanzee and gorilla occupy mainly western Africa, whereas ecological and paleontological evidence suggests proto-humans have migrated to eastern and southern Africa (Leakey et al. 2001). Hence a prolonged period of gene flow between ancestral chimpanzee and gorilla seems plausible.
Finally, we may analyze the joint effect of two speciation events in succession, as shown in Figure 1C. We assume that the species phylogeny of Figure 1C is strictly correct and the two allopatric events are separated by time t′ during which the effective population size was Ne′. The probability of having a genealogy incongruent with the species phylogeny, (m, M, M) or (M, m, M) of Figure 1C, is a function of t′/2Ne′ (Nei 1987; Wu 1991). The null hypothesis, again, is that t′/2Ne′ is the same for coding and intergenic regions. We used 53 intergenic and 76 coding sequences from human (H), chimpanzee (C), gorilla (G), and orangutan (O). Orangutan is used as an outgroup to distinguish the derived mutation, M, from the ancestral state, m. We masked all substitutions at CpG sites and then classified the patterns of independently segregating sites into the three categories shown in Figure 1C.
The proportion of incongruent genealogies is 0.509 and 0.361 for CDS and IGS, respectively (Table 3). The result of the likelihood-ratio test is not significant (P = 0.166), probably due to the small number of sequence fragments. With a larger sample size, say, twice the number of genes in Table 3, this approach should be useful for addressing the issue of allopatric speciation.
By analyzing the divergence among hundreds of DNA sequences, we inferred that the speciation history between human and chimpanzee cannot be the same for coding and intergenic regions. Genomic sequences between closely related species may thus provide new opportunities to settle the debate on the prevalence of allopatric speciation. In a series of analyses, Hey, Wakeley, and colleagues (Wakeley and Hey 1997; Kliman et al. 2000; Machado et al. 2002) addressed the same problem of parapatry using both the polymorphism and divergence data. While their approach utilizes more information per locus, we believe the approach outlined here will be more practical for several reasons. First, in the immediate future, there will be a torrent of data consisting of one sequence per gene per species. Second, polymorphism data will not be useful for resolving the mode of speciation in many species—human vs. chimpanzee being an obvious example. Third, the effect of selection on polymorphism can be more difficult to gauge than that on divergence, making the inference on speciation more difficult.
Finally, allopatric speciation could have more complex patterns than portrayed here. It may happen between deeply subdivided but connected populations where disparate genealogies preexisted when speciation took place allopatrically. Such a model can be seen as a hybrid between parapatry and allopatry. However, if populations can evolve to become differentially adapted and strongly subdivided in the presence of gene flow, it seems plausible that they can continue to diverge without a newly erected geographical barrier to stop gene flow completely. Moreover, the restriction of gene flow imposed by the diverging genomes should continue to strengthen as incompatibilities evolve to encompass larger and larger linkage blocks (Wu and Ting 2004). Testing such a hybrid model may require both the divergence and polymorphism data at the genomic level (Wakeley and Hey 1997; Kliman et al. 2000; Machado et al. 2002). At this moment, testing strict (and simple) allopatry among diverse taxa, as outlined here, seems a logical first step.
We thank T. Nagylaki, H. Tang, H. Y. Wang, J. Lu, and Y.-X. Fu for providing theoretical advice and/or helping with data analysis; F. C. Chen and W. H. Li for kindly providing the intergenic sequence data; R. Sakate and M. Hirai for the chimpanzee coding sequences; K. Hashimoto and C. K. J. Shen for the macaque cDNA sequences; and J. Shapiro, M. Kohn, B. Harr, M. Long, L. Zhang, I. Boussy, and J. Spofford for comments and discussions.
Communicating editor: Y.-X. Fu
- Received March 23, 2004.
- Accepted August 23, 2004.
- Genetics Society of America