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The look-ahead effect of phenotypic mutations

 




Dion J. Whitehead∗1, Claus O. Wilke∗2, David Vernazobres1, Erich Bornberg-Bauer1
( July 26, 2025)
Abstract

The evolution of complex molecular traits such as disulphide bridges often requires multiple mutations. The intermediate steps in such evolutionary trajectories are likely to be selectively neutral or deleterious. Therefore, large populations and long times may be required to evolve such traits. We propose that errors in transcription and translation may allow selection for the intermediate mutations if the final trait provides a large enough selective advantage. We test this hypothesis using a population based model of protein evolution. If an individual acquires one of two mutations needed for a novel trait, the second mutation can be introduced into the phenotype due to transcription and translation errors. If the novel trait is advantageous enough, the allele with only one mutation will spread through the population, even though the gene sequence does not yet code for the complete trait. The first mutation then has a higher frequency than expected without phenotypic mutations giving the second mutation a higher probability of fixation. Thus, errors allow protein sequences to ”look-ahead” for a more direct path to a complex trait.

Both authors contributed equally to this work.

1Institute for Evolution and Biodiversity, The Westphalian Wilhelms University of Muenster, 48149 Muenster, Germany

2Center for Computational Biology and Bioinformatics and Institute for Cell and Molecular Biology, Section of Integrative Biology, University of Texas, Austin, Texas 78712

Running Head: Look-ahead effect phenotypic mutations

Keywords: protein evolution, phenotype errors, mutation rates, complexity

Corresponding author:

Dion J. Whitehead

Institute for Evolution and Biodiversity,

The Westphalian Wilhelms University of Muenster,

Schlossplatz 4, D48149

Germany

Email: dion@uni-muenster.de

Phone: +49-(0)251-83-21633

Fax: +49-(0)251-83-21631

1 Introduction

According to a central principle of molecular evolution, the likelihood that a given mutation occurs is independent of the mutation’s phenotypic consequences. Organisms cannot choose specific mutations. This tenet was challenged by Cairns et al. (1988), who observed that under a certain selective pressure, E. coli cells appeared to acquire an excess of beneficial mutations. The idea that cells can somehow ‘direct’ evolution was thought provoking, and stimulated many investigations (for reviews see Bridges (1998); Foster (1999); Cairns (1998); Hall (1998); Rosenberg (2001)). While the notion that cells can directly decide in which genomic regions to increase their mutation rate has been mostly abandoned (Foster, 1998; Cairns, 1998), the original observations by Cairns et al. (1988) have been corroborated (see above reviews).

If mutations arise independently of their phenotypic consequences, then how can adaptations occur that require multiple amino acid mutations and for which the intermediate stages are either selectively neutral or disadvantageous? Large populations can climb multiple fitness peaks, even with disadvantageous intermediate alleles (Behe and Snoke, 2004; Weinreich and Chao, 2005). Although no new mechanisms are therefore required to explain the evolution of complex proteins (Lynch, 2005), we propose that errors in transcription and translation (phenotypic mutations) allow the selection of the intermediate mutations of a multiple-mutation requiring trait, and can thus speed up the evolution of complex traits.

Studies on the phenotypic mutation rate indicate that it is orders of magnitude larger than the genotypic mutation rate (Springgate and Loeb, 1975; Edelmann and Gallant, 1977): the global phenotypic non-synonymous mutation rate has been estimated to be 4.5×1044.5\times 10^{-4} mistranslations per codon (Ellis and Gallant, 1982; Shaw et al., 2002), compared with a genotypic mutation rate of between 107\sim 10^{-7} to 101110^{-11} (Drake et al., 1998). Consequently, for a protein of 300 residues, on average more than 1 in 10 copies of the protein will contain a mutation. Using mutation rates derived from the literature and conservative biological assumptions, we show via mathematical modeling and simulations that phenotypic mutations allow evolution to select for neutral intermediate alleles of a multi-mutation trait, actually selecting for proteins whose exact DNA sequence is not in the organism under selection. Evolution is then able to look ahead for evolutionary jackpots in sequence space.

Our theory is based on the following hypothetical scenario. A protein can increase the fitness of an individual if it evolved a specific trait. This trait requires two mutations, for example a disulphide bridge between two cysteine residues. Having only one of the required mutations is either selectively neutral or deleterious, however when an individual has only one mutation, small amounts of the protein with both mutations will be produced due to phenotypic mutations. If the presence of both mutations at low concentrations provides even a small fitness improvement then the allele with one mutation will spread though the population. As the frequency of the intermediate allele increases, there is a greater probability that if the second mutation occurs, it will be the presence of the first mutation, and thus provide the full fitness benefit.

The aim of this article is to show that adaptive phenotypic mutations can undergo positive selection under biologically plausible conditions, allowing proteins greater access to features involving multiple mutations. We thus introduce the notion of a ”look-ahead” effect. The name indicates that seemingly unsurmountable evolutionary barriers can be overcome thanks to phenotypic mutations which are not yet present in the genome. We wish to emphasize that the look-ahead concept is firmly grounded on the idea of chance and necessity and by no means insinuates a teleological feature of molecular evolution.

2 Materials and Methods

Variables used in this work
β\beta Phenotypic mutation rate from allele 1 to allele 2
μ\mu Null mutation rate per residue
NN Population size
rr Number of residues available for the second mutation
ss Selection coefficient for allele 2
UU Generational (DNA) mutation rate from allele 1 to allele 2

Model assumptions:

We model the scenario of a protein evolving a trait that requires two mutations. The model is based on a population-genetics framework where a single gene can evolve into different alleles. We do not consider duplication and divergence of genes. In addition, the process described here will likely only occur for proteins with sufficiently long half-lives, as the protein must persist for some time to exert a phenotypic effect. As we model only a single gene, we expect our results to be more relevant for single-celled organisms and viruses than for multicellular organisms, which tend to have larger genomes and smaller effective population sizes than microorganisms.

Refer to caption
Figure 1: The three alleles (or genotypes). The vertical lines in the genes indicate the number of key mutations required for the novel two-residue function. The fitness of the allele 1 increases if phenotypic mutations are taken into consideration.

The model consists of the evolution of three non-recombining haploid genotypes, where each genotype contains one of the three alleles shown in Figure 1. The three different alleles are named according to number of relevant mutations, corresponding to zero mutations (allele 0), a single mutation (allele 1), and both mutations (allele 2) required for the adaptive feature. Having both mutations of the adaptive feature provides a selective advantage ss. We assume that the intermediate allele (allele 1) is selectively neutral if transcribed and translated without error. We specifically take into consideration errors in transcription and translation, that is, phenotypic mutations.

In the model, the population initially consists of one individual carrying allele 1 and N1N-1 individuals carrying allele 0. So long as allele 1 is present, allele 2 can be generated by mutations. The population evolves for a fixed time period, during which allele 2 can be generated by mutation and go to fixation.

In each generation, selection increases the frequency of the alleles according to their corresponding fitness values. Allele 0 has a fitness of 1. Allele 2 has a fitness of 1+s1+s, where ss is the selection coefficient provided by the adaptive feature. The fitness of allele 1, the intermediate allele with only a single mutation, depends on the phenotypic error rate. Most phenotypic errors will be neutral or deleterious, however some will be beneficial. For simplicity, we assume that the length of the protein and the expression level are both constant. In addition, we also assume that the cost due to deleterious phenotypic errors is also constant. The effect of these parameters on this model will be the subject of future work. If there are no phenotypic mutations, allele 1 has the same fitness as allele 0. However, if phenotypic mutations occur, allele 1 can produce a small number of allele 2 proteins due to phenotypic errors. The fitness of allele 1 is therefore dependent on the number of such errors.

We assume that fitness is a linear sum of individual proteins, meaning that if some phenotypic variants of a protein have a higher fitness, then the overall fitness of that allele is proportionally increased.

We let rr be the number of residues that can potentially complement the first mutation to provide the full two-residue adaptive feature. These rr residues represent, e.g., the sites at which the second cysteine of a cysteine bridge could arise; other similar two-residue mutations that significantly improve functionality can be proposed. Two residues that comprise an adaptive trait are likely to co-evolve, because if a mutation occurs in one of the residues, selection strongly favors a compensatory mutation in the other. Based on a large data set, Martin et al. (2005) found that co-evolving residues are spatially near. Co-evolving residues were, on average, 98.6 amino-acids apart along the sequence, but had a mean spatial distance of 6.9 Å. This spatial distance can be compared to the width of the van der Waals volume of an amino-acid (5-6 Å), showing that most co-evolving residues are effectively in contact proximity. Therefore, rr is mostly independent of the size of the protein, as long as the protein is of sufficient length. Bloom et al. (2006) calculated the mean contact density (the mean number of residues in contact with a given residue) for 194 yeast proteins, and found that most residues have a mean contact density of seven to eight residues. In this work we use r=8r=8.

Given rr possible positions for the second residue, and assuming that each position requires a specific residue, the fraction of proteins of allele 1 containing the second (now highly beneficial) mutation is β=r19λ\beta=\frac{r}{19}\lambda, where λ\lambda is the per codon non-synonymous phenotypic mutation rate. In this model, we use λ=4.5×104\lambda=4.5\times 10^{-4} mistranslations per codon (Ellis and Gallant, 1982; Shaw et al., 2002). The fraction β\beta of allele-2 proteins contribute to the fitness, giving allele 1 a fitness of 1+sβ1+s\beta.

When considering genetic (i.e. inherited) mutations, for simplicity we neglect back mutations (e.g. from allele 1 to allele 0), and assume there are no recurrent mutations of allele 1 from allele 0 (the model starts with a single copy of allele 1). Allele 2 arises via a mutation from allele 1. We ignore the possibility of a double mutation directly from allele 0 to allele 2, as this probability is extremely small. The genetic mutation rate for allele 1 mutating into allele 2 is derived as follows: For microbes, the rate of mutations per nucleotide per generation is between 107\sim 10^{-7} to 101110^{-11} (Drake et al., 1998). Here we use 10810^{-8} as the non-synonymous mutation rate per codon per generation. The resulting mutation rate for changing allele 1 into allele 2 is U=r19108=819108U=\frac{r}{19}10^{-8}=\frac{8}{19}10^{-8}.

Genes can also acquire null mutations, rendering the gene non-functional and therefore eliminating the organism. The null mutation rate for protein-encoding genes is on the order of 10610^{-6} per generation (Drake et al., 1998). However, this rate will depend on the length (LL) of the protein. Assuming an average protein length of 300 residues, the per-residue null mutation rate is given by 106/300=3.3910^{-6}/300=\sim 3.3^{-9}. For a protein of length LL, the null mutation rate is given by μ=3.39L\mu=3.3^{-9}L.

Simulations:

The numerical simulations were written in Java using the Colt scientific library (Colt Project, 2007) for the generation of random numbers. The analytic expressions were evaluated using both Mathematica and Python, the latter in conjunction with the SciPy package (SciPy, 2007). Source code for the numerical simulations is available on request from DJW.

The population in each simulation is represented by three numbers, corresponding to the abundance of each of the three alleles. As described, the initial abundances are N1,1,0N-1,1,0 for alleles 0,1,20,1,2, respectively. The simulation runs for a specified number of generations TT. We used T=5×105T=5\times 10^{5} throughout this work. Strictly speaking, TT is the number of generations in which allele 1 can mutate into allele 2; for later generations this possibility of mutation is disabled. If allele 2 is present at time TT, then the simulation is continued until allele 2 is either lost or has reached fixation. Generations are discrete, with mutations, selection, and drift occurring at each generation. During each generation we perform the following steps. First we check if either allele 0 or allele 2 has reached fixation; if so, we stop the simulation, as both cases are absorbing states. Next, for each allele we check for null mutations by drawing a random number from the Poisson distribution where the expected number of events is the null mutation rate μ\mu multiplied by the total number of individuals with the given allele. Mutations from allele 1 to allele 2 are computed in a similar manner, where the expected number of events is UU multiplied by the number of allele 1 individuals. Then, after the possible production of the mutant allele 2, selection acts on the fitness of the alleles, where the frequency of each allele is multiplied by its corresponding fitness, [1,1+sβ,1+s][1,1+s\beta,1+s] for alleles [0,1,2][0,1,2], giving the new number of alleles in a possibly larger population. Finally, the next population of NN individuals is chosen by recursively sampling from the binomial distribution, representing random genetic drift. Allele 0 is first sampled with the mean=(frequency of allele 0), and the (number of trials)=NN. Allele 1 is then sampled from the combined allele 1 and 2 individuals. The number of simulations where allele 2 becomes fixed is divided by the total number of simulations, giving an estimate of the fixation probability. The number of simulations for each parameter set was between 10810^{8} and 10910^{9}.

3 Results

3.1 Analytical fixation rate of allele 2

To calculate the fixation rate of allele 2 we have to consider the two fates of allele 1. Firstly, allele 1 can become lost. In this case allele 2 can only be generated during the period of drift of allele 1. The alternative fate of allele 1 is fixation. Then allele 2 can be generated either while allele 1 drifts to fixation or after allele 1 is already fixed. We would like to know how many mutation events from allele 1 to allele 2 are expected for either fate of allele 1. We let n(sβ)n(s\beta) be the expected number of mutation events for when allele 1 is eventually fixed, and nloss(sβ)n_{loss}(s\beta) be the expected number of mutation events for the case when allele 1 is lost. We can calculate n(sβ)n(s\beta) and nloss(sβ)n_{loss}(s\beta) from diffusion theory, by integrating over the sojourn times of allele 1. The corresponding calculations are cumbersome but straightforward, and for the sake of brevity we present the details in the Appendix (A.4 and A.5). For n(sβ)n(s\beta), allele 2 can be generated as allele 1 drifts to fixation, and also after allele 1 has already reached fixation. For nloss(sβ)n_{loss}(s\beta), allele 2 can only be generated while allele 1 drifts.

Assuming that mm is the expected number of times allele 2 is generated, what is the probability that at least one copy goes to fixation? The probability of fixation of a single copy of allele 2 is u(s)u(s) Kimura (1962). (In Appendix A.1, we reproduce the exact expression for u(s)u(s), as well as approximations for large and small ss.) Thus, if allele 2 is generated kk times, its probability of fixation is 1[1u(s)]k1-[1-u(s)]^{k}. Since the probability that allele 2 is generated kk times follows a Poisson distribution with mean mm, we find for the probability vv that at least one of the mutations to allele 2 goes to fixation

v\displaystyle v =1kmkk!em[1u(s)]k\displaystyle=1-\sum_{k}\frac{m^{k}}{k!}e^{-m}[1-u(s)]^{k}
=1emu(s).\displaystyle=1-e^{-mu(s)}. (1)

We calculate this probability separately for n(sβ)n(s\beta) and nloss(sβ)n_{loss}(s\beta), setting mm equal to either of these values. We assume that TT is sufficiently large so that allele 1 has time to reach fixation within this interval (we assume T2NT\gtrsim 2N). Then the probability u2(s,β)u_{2}(s,\beta) that allele 2 is generated and goes to fixation (starting with a single copy of allele 1) is

u2(s,β)=u(sβ)(1en(sβ)u(s))+(1u(sβ))(1enloss(sβ)u(s)).u_{2}(s,\beta)=u(s\beta)\big{(}1-e^{-n(s\beta)u(s)}\big{)}+\big{(}1-u(s\beta)\big{)}\big{(}1-e^{-n_{\text{loss}}(s\beta)u(s)}\big{)}. (2)

The first half of the equation stems from the case when allele 1 eventually reaches fixation, where the probability that allele 1 becomes fixed, u(sβ)u(s\beta), is multiplied by the probability vv that at least one copy of allele 2 is generated and fixed. The second half corresponds to the case of loss of allele 1 from the population, where the probability of loss of allele 1, (1u(sβ))(1-u(s\beta)), is multiplied by the probability of at least one mutation from allele 1 to allele 2 and subsequent fixation of allele 2. Taking into account allele 2 mutations during allele 1 loss is important especially for small ss. Allele 1 is more likely to be lost than fixed for small ss, but can occasionally drift for long times before being lost.

In the limit β0\beta\rightarrow 0, i.e., in the absence of phenotypic mutations, we find with Eqs. (A2), (A27), and (A35)

u2(s,0)=N+1NeNU(TN)u(s)/NeNUu(s).u_{2}(s,0)=\frac{N+1}{N}-e^{-NU(T-N)u(s)}/N-e^{-NUu(s)}. (3)

(We assume that N1N\gg 1, and neglect corrections of order 1 compared to NN. Note that we cannot simplify (N+1)/N(N+1)/N to 1, because for small UU, 1eNUu(s)1-e^{-NUu(s)} and (1eNU(TN)u(s))/N(1-e^{-NU(T-N)u(s)})/N are of the same order in NN.)

As we are interested in the effect of phenotypic mutations (β>0\beta>0) compared to the case without phenotypic mutations (β=0\beta=0), we define the increase in the probability of fixation from advantagous phenotypic mutations (the look-ahead effect) as

ξ=u2(s,β)u2(s,0).\xi=\frac{u_{2}(s,\beta)}{u_{2}(s,0)}. (4)

We can broaden the assumption of T2NT\gtrsim 2N to TT\rightarrow\infty with good accuracy. For TT\rightarrow\infty, if allele 1 is destined to reach fixation, then the probability of generating at least one copy of allele 2 that goes to fixation approaches 1. Therefore, 1en(sβ)u(s)11-e^{-n(s\beta)u(s)}\rightarrow 1, in this limit, and thus

ξu(sβ)+(1u(sβ))(1enloss(sβ)u(s))(N+1)/NeNUu(s).\xi\approx\frac{\displaystyle u(s\beta)+\big{(}1-u(s\beta)\big{)}\big{(}1-e^{-n_{\text{loss}}(s\beta)u(s)}\big{)}}{(N+1)/N-e^{-NUu(s)}}. (5)

Apart from a correction for the case when allele 2 occurs while allele 1 is destined for extinction, Equation (5) is just the ratio of the probability of allele-1 fixation in the presence and absence of phenotypic mutations, u(sβ)/u(0)=Nu(sβ)u(s\beta)/u(0)=Nu(s\beta).

To first order in sβs\beta, Eq. (5) simplifies to (Appendix A.6)

ξ1+Nsβ+𝒪(s2β2).\xi\approx 1+Ns\beta+{\cal O}(s^{2}\beta^{2}). (6)

We can see from this equation that the look-ahead effect becomes important when NN is on the order of 1/(sβ)1/(s\beta).

For Nsβ1Ns\beta\gg 1, only the first term contributes to the numerator in Eq. (5), and we obtain (Appendix A.7)

ξ(1e2sβ)(N+1)/Nexp[NU(1e2s)].\xi\approx\frac{\displaystyle(1-e^{-2s\beta})}{\displaystyle(N+1)/N-\exp[-NU(1-e^{-2s})]}. (7)

3.2 Simulations

Refer to caption
Figure 2: Fixation probability of allele 2 (u2u_{2}) vs. the selection coefficient ss. Black is for u2(s,β)u_{2}(s,\beta), grey is for u2(s,0)u_{2}(s,0). Solid lines are predictions according to Eq. (2) and (3), data points are for simulations with 10910^{9} repeats. N=104N=10^{4}, U=819108U=\frac{8}{19}10^{-8}, β=0.00019\beta=0.00019, T=5×105T=5\times 10^{5}. Error bars are standard errors.

We confirmed our analytic results for the fixation probabilities u2(s,β)u_{2}(s,\beta) and u2(s,0)u_{2}(s,0) by numerical simulation, for different values of ss (Figure 2). With a population size N=104N=10^{4}, the effect of phenotypic mutations can be seen for s>0.1s>0.1, and increases for larger ss. For s<0.1s<0.1, the effect is too small and the intermediate allele is effectively neutral, meaning the fixation of allele 2 depends on the random fixation of the neutral allele 1. The look-ahead effect, ξ\xi, shows the simulation results compared to Equations (5), (6) and (7). Figure 3 shows the magnitude of the look-ahead effect for the same parameter settings. For large ss, the look-ahead effect can inflate the probability of fixation of allele 2 by several orders of magnitude.

Refer to caption
Figure 3: Look-ahead effect (ξ\xi) due to phenotypic mutations vs. the selection coefficient ss. The solid line is for Eq. (5), dashes are for Eq. (6), dots are for Eq. (7), and data points are for simulations with 10910^{9} repeats. N=104N=10^{4}, U=819108U=\frac{8}{19}10^{-8}, β=0.00019\beta=0.00019, T=5×105T=5\times 10^{5}. Error bars are standard errors.

We also display the different analytic expressions for ξ\xi in Figure 3. The approximation (5), derived in the limit TT\rightarrow\infty, works well for all values of ss. The approximation (6), derived for small sβs\beta, captures correctly the magnitude of ss at which the look-ahead effect starts to operate, i.e., s1/(Nβ)s\gtrsim 1/(N\beta). Similarly, approximation (7), valid for Nsβ1Ns\beta\gg 1, approximates ξ\xi well for larger ss.

Figure 4 shows ξ\xi for different population sizes. As expected from the condition s1/(Nβ)s\gtrsim 1/(N\beta), the look-ahead effect will work with smaller selection coefficients ss in larger populations. For large ss, ξ\xi saturates at approximately NN.

Refer to caption
Figure 4: Look-ahead effect (ξ\xi) due to phenotypic mutations vs. the selection coefficient ss for different population sizes (NN). Solid lines are from Equation (5), data points are for simulations with 10810^{8} repeats. U=819108U=\frac{8}{19}10^{-8}, β=0.00019\beta=0.00019, T=5×105T=5\times 10^{5}. Error bars are standard errors.

We studied the effect of different values of the phenotypic error rate β\beta (Fig. 5). As the error rate β\beta increases, the look-ahead effect ξ\xi increases by the same order of magnitude. For a very high phenotypic error rate of β=0.019\beta=0.019, the look-ahead effect is present for very small values of ss. However, such a high error rate is likely to be severely detrimental, and in our model we do not take into account the loss of overall fitness for increasing phenotypic error rates. Conversely for smaller β\beta, the look-ahead effect is restricted to large ss.

Refer to caption
Figure 5: Look-ahead effect (ξ\xi) due to phenotypic mutations vs. the selection coefficient ss for different phenotypic error rates (β\beta). Solid lines are from Equation (5), data points are for simulations with 10810^{8} repeats. N=104N=10^{4}, U=819108U=\frac{8}{19}10^{-8}, T=5×105T=5\times 10^{5}. Error bars are standard errors.

4 Discussion

We have described a model demonstrating the consequences of positive phenotypic mutations on the evolution of a single gene. We have compared numerical simulations with the analytical approximations and found them to be in good agreement. When phenotypic mutations exert an effect on fitness, selection can operate on the intermediate allele of a complex trait, which otherwise (without phenotypic mutations) would be neutral. We refer to selection for the intermediate allele as the look-ahead effect, because this effect allows evolution to select for sequences not yet in the genome.

The approximation for small sβs\beta, Eq.(6), shows most clearly the relationship between the parameters. The look-ahead effect is proportional to NN, ss, and β\beta, and sets in when NN is on the order of 1/(sβ)1/(s\beta). For large NsβNs\beta, the look-ahead effect saturates. The asymptotic value of ξ\xi is approximately NN for NU1NU\ll 1. Therefore, large populations have two advantages over small populations in terms of the look-ahead effect: the effect sets in for smaller values of ss, and saturates at a larger asymptotic value ξ\xi. Of course, even in the absence of the look-ahead effect, larger populations can more easily traverse multiple local fitness peaks (Weinreich and Chao, 2005).

Because the selection coefficient ss depends on the environment, a valid question is how often does ss reach sufficiently high levels so that the look-ahead effect can operate. For microbial species such as bacteria, sufficiently large ss should be reasonably common. Many bacteria experience highly fluctuating (Smit et al., 2001) and structured (Baquero and Negri, 1997) environments, where growth is limited by the lack of a key trait. An obvious and extreme example is antibiotic resistance. Evolving a defense against an antibiotic molecule can involve only a few amino acids (Palzkill et al., 1994), and those individuals that can generate an enzyme capable of degrading the antibiotic, even if briefly or weakly, will have a very large fitness increase. In fact, if the efficacy of the antibiotic is 100% on susceptible genotypes, a mutation providing only moderate resistance has an infinite selective advantage. And even for very small antibiotic concentrations, mutants differing by only two amino acids at a single β\beta-lactamase gene can be selected effectively (Baquero et al., 1997, 1998). Thus, bacteria may frequently experience environments in which a large fitness increase (large ss) is only a few mutations away. Similarly, in bacteriophages, selective coefficients ss of 10 or more are not uncommon, even for individual mutations (Bull et al., 2000).

Our work is entirely theoretical, but we expect that it will be possible to experimentally verify our predictions in future work. For experimentally observing the look-ahead effect, we would need a system where ss and NN are both large, while β\beta (the phenotypic mutation rate) can be modified. The values of both NN and ss used in this work are well within biologically realistic ranges achievable in a microbiological laboratory. Conditions for large ss may be created with e.g. antibiotic resistance, which is a common laboratory workhorse. Unfortunately, many antibiotics function by reducing translation fidelity (Ogle and Ramakrishnan, 2005), and thus would conflate ss and β\beta. Changing β\beta could involve a mutated ribosome. Ribosomes appear to be optimized for accurate and efficient translation of mRNA (Baxter-Roshek et al., 2007), and several examples of altered ribosome fidelity exist, both increasing (Vila-Sanjurjo et al., 2003) and decreasing fidelity (Friedman et al., 1968). Specific regions of the ribosome rRNA sequence have been identified as influencing fidelity (O’Connor et al., 1997), and various agents can reduce fidelity, e.g., streptomycin, magnesium, and ethanol (Friedman et al., 1968). Few mutations may be sufficient to alter the fidelity of a ribosome, for example, a single mutation in the S5 ribosomal protein in E. coli increases frameshifting and nonsense mutations (Kirthi et al., 2006). In yeast, mutations in the 18S RNA have been found that both increase and decrease translational fidelity (Konstantinidis et al., 2006).

In this work, we have calculated the look-ahead effect from a comparison between the two cases of β>0\beta>0 and β=0\beta=0. The latter may not be experimentally possible; any experiment will likely compare two different positive values of β\beta. Nevertheless, Figure 5 shows that a larger look-ahead effect can be achieved with a higher β\beta, where increasing β\beta by one order of magnitude both increases the look-ahead effect by an order of magnitude and lowers the smallest ss where an effect is observed. Of course, our model does not take into account the loss of fitness or other confounding effects from a higher phenotypic mutation rate. Thus, a balance must be found in having two different values of β\beta that are different enough to measure, while at the same time minimizing the confounding effects. The most obvious consequence of increasing the phenotypic mutation rate is that overall fitness may be reduced, for example in E. coli, where a higher translational error rate activates stress responses (Fredriksson et al., 2007), or in mouse, where such errors are implicated in neurodegeneration caused by misfolded proteins that aggregate (Lee et al., 2006). Increasing translational fidelity may not come without fitness cost either. The hyperaccurate mutations in the 18S RNA in yeast (Konstantinidis et al., 2006) cause an increase in oxidative stress. This observation suggests that cells consume more energy to achieve hyperaccuracy. It may also partially explain why the phenotypic error rate is much higher than the genotypic error rate, as there is possibly a direct disadvantage in reducing the phenotypic error rate, rather than only reducing the selective advantage that occurs if the phenotypic error rate is reduced, as discussed in Buerger et al. (2006).

Buerger et al. (2006) asked whether evolution has selected for the current phenotypic error rate, which does not differ significantly between eukaryotes and prokaryotes (Ellis and Gallant, 1982; Shaw et al., 2002) even though the source of errors is different. They suggested that the increase in fitness becomes incrementally smaller for improvements to transcription and translation fidelity. We would like to speculate that the phenotypic error rate is on the border between minimal costs (of e.g. misfolded proteins) and maximum payoff (via the look-ahead effect).

The goal of our analysis was to demonstrate that the look-ahead effect is theoretically possible, and as such, we intentionally excluded confounding factors for the sake of clarity. There are several aspects not considered in our model that may play important roles. For example, in this work we did not consider the expression level. For low expressed genes, the mutation from allele 1 to allele 2 will occur less frequently compared to highly expressed genes. However, if allele 2 is produced it will be at a higher concentration (of allele 2 mutant proteins in a population of allele 1 proteins), as the overall copy number of allele 1 is low. This difference in expression levels is likely reduced in a large population, where beneficial mutations occur with sufficient frequency. Another factor related to the expression level is translational robustness. It has been proposed that highly expressed genes are under selection to properly fold despite phenotypic mutations, and consequently evolve slower (Drummond et al., 2005; Wilke and Drummond, 2006). If a gene is robust to translational errors, then it can tolerate a larger variety of mutations, of which some may be intermediates to a new adaptive multi-residue trait. Thus, translational robustness may increase the sequences available for experimentation at the phenotypic level. However, if the intermediate allele is itself not robust to errors in translation, then it will not be neutral, and may be selected against. The location of the protein trait will also influence the viability of the intermediate allele: mutations near the surface of the protein are less likely to disrupt the protein compared to mutations in the core (Tokuriki et al., 2007).

In conclusion, we propose that organisms can experiment with protein sequences that are mutationally close to the current sequence, but not yet in the genome. This effect allows selection for intermediates of complex traits, opening up a more direct route to the trait and thus reducing the time needed for fixation in the population.

5 Acknowledgements

D.J.W. would like to thank January Weiner for stimulating discussions and Maya Amago for helpful suggestions. D.J.W. and E.B.B. were supported by an HFSP program grant. C.O.W. was supported by NIH grant AI 065960.

Appendix A Appendix

Here, we present the details of our analytic derivations.

A.1 Probability of fixation

According to (Kimura, 1962), the probability of fixation u(s)u(s) of a single allele with selection coefficient ss is given by

u(s)=1e2s1e2Ns.u(s)=\frac{1-e^{-2s}}{1-e^{-2Ns}}. (A1)

For s1/Ns\lesssim 1/N, this expression simplifies to

u(s)=1N+N1Ns+𝒪(s2),u(s)=\frac{1}{N}+\frac{N-1}{N}s+{\cal O}(s^{2}), (A2)

whereas for Ns1Ns\gg 1, this expression simplifies to

u(s)1e2s.u(s)\approx 1-e^{-2s}. (A3)

A.2 A single allele drifting to fixation or loss

We first consider a single allele with selective advantage ss drifting to fixation or extinction, and ask how many mutations this allele generates until it is either fixed or lost. We will treat these two cases separately. Let nfix(s)n_{\text{fix}}(s) be the expected number of mutations generated while the allele drifts to fixation, and let nloss(s)n_{\text{loss}}(s) be the expected number of mutations generated while the allele drifts to extinction. We calculate these two quantities using diffusion theory, by integrating the sojourn times of the allele over all frequencies.

For an allele with selective coefficient ss and starting at frequency p=1/Np=1/N, Nagylaki (1974) calculated its mean sojourn time τ(y)\tau(y) between frequencies yy and y+dyy+dy as

τ(y)=2[V(y)G(y)]1[uloss(1/N)g(0,y)θ(1/Ny)+ufix(1/N)g(y,1)θ(y1/N)].\tau(y)=2[V(y)G(y)]^{-1}[u_{\text{loss}}(1/N)g(0,y)\theta(1/N-y)+u_{\text{fix}}(1/N)g(y,1)\theta(y-1/N)]. (A4)

Here,

V(y)G(y)\displaystyle V(y)G(y) =y(1y)e2Nsy/N,\displaystyle=y(1-y)e^{-2Nsy}/N, (A5)
g(a,b)\displaystyle g(a,b) =e2Nsae2Nsb2Ns,\displaystyle=\frac{e^{-2Nsa}-e^{-2Nsb}}{2Ns}, (A6)
uloss(p)\displaystyle u_{\text{loss}}(p) =e2Nspe2Ns1e2Ns,\displaystyle=\frac{e^{-2Nsp}-e^{-2Ns}}{1-e^{-2Ns}}, (A7)
ufix(p)\displaystyle u_{\text{fix}}(p) =1uloss(p)=1e2Nsp1e2Ns,\displaystyle=1-u_{\text{loss}}(p)=\frac{1-e^{-2Nsp}}{1-e^{-2Ns}}, (A8)

and θ(z)\theta(z) is the Heaviside step function. We want to integrate expressions involving τ(y)\tau(y) from y=0y=0 to y=1y=1. Since y=1/Ny=1/N corresponds to a single copy of the allele that drifts to fixation, values of yy less than 1/N1/N are not relevant for our analysis. Therefore, we discard the term proportional to θ(1/Ny)\theta(1/N-y) in Eq. (A4), and use in what follows

τ(y)=2ufix(1/N)g(y,1)/[V(y)G(y)]for y>1/N.\tau(y)=2u_{\text{fix}}(1/N)g(y,1)/[V(y)G(y)]\qquad\text{for $y>1/N$}. (A9)

A.3 Number of mutations conditional on fixation

For the sojourn time conditional on fixation, τfix(y)\tau_{\text{fix}}(y), Nagylaki (1974) finds

τfix(y)=τ(y)ufix(y)/ufix(p).\tau_{\text{fix}}(y)=\tau(y)u_{\text{fix}}(y)/u_{\text{fix}}(p). (A10)

Using this expression, we have

nfix(s)=NU1/N1τfix(y)y𝑑y.n_{\text{fix}}(s)=NU\int_{1/N}^{1}\tau_{\text{fix}}(y)ydy. (A11)

Plugging the expressions for V(y)G(y)V(y)G(y), g(a,b)g(a,b), ufix(p)u_{\text{fix}}(p), and τ(y)\tau(y) into τfix(y)\tau_{\text{fix}}(y), we arrive at

τfix(y)=1s(1e2Ns)(1e2Nsy)(1e2Ns(1y))y(1y).\tau_{\text{fix}}(y)=\frac{1}{s(1-e^{-2Ns})}\frac{(1-e^{-2Nsy})(1-e^{-2Ns(1-y)})}{y(1-y)}. (A12)

This expression corresponds to the one by Ewens (1973). Note that yτfix(y)0y\tau_{\text{fix}}(y)\rightarrow 0 for y0y\rightarrow 0. Therefore, we can extend the lower limit of integration to 0 in Eq. (A11), and rewrite nfix(s)n_{\text{fix}}(s) as

nfix(s)=NUs(1e2Ns)I(2Ns)n_{\text{fix}}(s)=\frac{NU}{s(1-e^{-2Ns})}I(2Ns) (A13)

with

I(a)=01(1eay)(1ea(1y))1y𝑑y.I(a)=\int_{0}^{1}\frac{(1-e^{-ay})(1-e^{-a(1-y)})}{1-y}dy. (A14)

The integral I(a)I(a) can be rewritten as

I(a)=γEi(a)+ln(a)+ea[γEi(a)+ln(a)],I(a)=\gamma-\text{Ei}(-a)+\ln(a)+e^{-a}[\gamma-\text{Ei}(a)+\ln(a)], (A15)

where γ0.5772\gamma\approx 0.5772 is the Euler-Mascheroni constant and Ei(z)\text{Ei}(z) is the exponential integral,

Ei(z)=zett𝑑t.\text{Ei}(z)=-\int_{-z}^{\infty}\frac{e^{-t}}{t}dt. (A16)

For s1/Ns\lesssim 1/N, we find

nfix(s)=N2U+𝒪(s2).n_{\text{fix}}(s)=N^{2}U+{\cal O}(s^{2}). (A17)

For Ns1Ns\gg 1, we obtain the asymptotic expansion

nfix(s)NUs[ln(2Ns)+γ],n_{\text{fix}}(s)\approx\frac{NU}{s}[\ln(2Ns)+\gamma], (A18)

using Abramowitz and Stegun (1964) 5.1.51,

Ei(z)ezz(11z+2z26z3)for large z.\text{Ei}(-z)\sim-\frac{e^{-z}}{z}\Big{(}1-\frac{1}{z}+\frac{2}{z^{2}}-\frac{6}{z^{3}}\Big{)}\qquad\text{for large $z$}. (A19)

A.4 Number of mutations conditional on extinction

For the sojourn time conditional on extinction, τloss(y)\tau_{\text{loss}}(y), Nagylaki (1974) finds

τloss(y)=τ(y)uloss(y)/uloss(p).\tau_{\text{loss}}(y)=\tau(y)u_{\text{loss}}(y)/u_{\text{loss}}(p). (A20)

Using this expression, we have

nloss(s)=NU1/N1τloss(y)y𝑑y.n_{\text{loss}}(s)=NU\int_{1/N}^{1}\tau_{\text{loss}}(y)ydy. (A21)

Plugging the expressions for V(y)G(y)V(y)G(y), g(a,b)g(a,b), uloss(p)u_{\text{loss}}(p), and τ(y)\tau(y) into τloss(y)\tau_{\text{loss}}(y), we find

τloss(y)=1s(1e2Ns)e2s11e2(N1)s(e2Nsye2Ns)(1e2Ns(1y))y(1y).\tau_{\text{loss}}(y)=\frac{1}{s(1-e^{-2Ns})}\frac{e^{2s}-1}{1-e^{-2(N-1)s}}\frac{(e^{-2Nsy}-e^{-2Ns})(1-e^{-2Ns(1-y)})}{y(1-y)}. (A22)

We rewrite nlossn_{\text{loss}} as

nloss=NUs(1e2Ns)e2s11e2(N1)sJ(N,s)n_{\text{loss}}=\frac{NU}{s(1-e^{-2Ns})}\frac{e^{2s}-1}{1-e^{-2(N-1)s}}J(N,s) (A23)

with

J(N,s)=1/N1(e2Nsye2Ns)(1e2Ns(1y))1y𝑑y.J(N,s)=\int_{1/N}^{1}\frac{(e^{-2Nsy}-e^{-2Ns})(1-e^{-2Ns(1-y)})}{1-y}dy. (A24)

The integral can be rewritten as

J(N,s)=2e2Ns(γChi[2(N1)s]+ln[2(N1)s]),J(N,s)=-2e^{-2Ns}\big{(}\gamma-\text{Chi}[2(N-1)s]+\ln[2(N-1)s]\big{)}, (A25)

where Chi(z)\text{Chi}(z) is the hyperbolic cosine integral,

Chi(z)=γ+ln(z)+0zcosh(t)1t𝑑t.\text{Chi}(z)=\gamma+\ln(z)+\int_{0}^{z}\frac{\cosh(t)-1}{t}dt. (A26)

For s1/Ns\lesssim 1/N, we find

nloss(s)=(N1)U+𝒪(s2).n_{\text{loss}}(s)=(N-1)U+{\cal O}(s^{2}). (A27)

For Ns1Ns\gg 1, we obtain the asymptotic expansion

nloss(s)U2s2(1e2s),n_{\text{loss}}(s)\approx\frac{U}{2s^{2}}(1-e^{-2s}), (A28)

using

Chi(z)Ei(z)2ez2zfor large z.\text{Chi}(z)\approx\frac{\text{Ei}(z)}{2}\approx\frac{e^{z}}{2z}\qquad\text{for large $z$.} (A29)

[This expansion follows directly from the definitions of Chi(z)\text{Chi}(z), cosh(z)\cosh(z), and Ei(z)\text{Ei}(z).]

A.5 Number of mutations within a given time interval

We now extend the derivations in Section A.3 to calculate the number of mutations to allele 2 generated within a certain time interval TT, conditional on fixation of allele 1. We assume that TT is sufficiently large so that allele 1 has time to reach fixation within this interval. We only consider the case conditional on fixation because no new mutations are generated once allele 1 has gone extinct.

We calculate n(s)=nfix(s)+nT(s)n(s)=n_{\text{fix}}(s)+n_{\text{T}}(s), where nT(s)n_{\text{T}}(s) is the total number of mutations generated once the first mutation has reached fixation. We have

nT(s)=NU[Ttfix(s)],n_{\text{T}}(s)=NU[T-t_{\text{fix}}(s)], (A30)

where tfix(s)t_{\text{fix}}(s) is the time to fixation of a mutation with selective advantage ss. This time is given by the integral over all sojourn times,

tfix(s)=01τfix(y)𝑑y=I2(2Ns)s(1e2Ns)t_{\text{fix}}(s)=\int_{0}^{1}\tau_{\text{fix}}(y)dy=\frac{I_{2}(2Ns)}{s(1-e^{-2Ns})} (A31)

with

I2(a)=01(1eay)(1ea(1y))y(1y)𝑑y.I_{2}(a)=\int_{0}^{1}\frac{(1-e^{-ay})(1-e^{-a(1-y)})}{y(1-y)}dy. (A32)

A partial fraction decomposition of the integrand reveals that I2(a)=2I(a)I_{2}(a)=2I(a), and thus we have

tfix(s)=2I(2Ns)s(1e2Ns)t_{\text{fix}}(s)=\frac{2I(2Ns)}{s(1-e^{-2Ns})} (A33)

Combining this result with Eqs. (A13) and (A30), we find

n(s)\displaystyle n(s) =nfix(s)+nT(s)=NU[TI(2Ns)s(1e2Ns)]\displaystyle=n_{\text{fix}}(s)+n_{\text{T}}(s)=NU\Big{[}T-\frac{I(2Ns)}{s(1-e^{-2Ns})}\Big{]}
=NUTnfix(s).\displaystyle=NUT-n_{\text{fix}}(s). (A34)

Note that n(s)=nfix(s)n(s)=n_{\text{fix}}(s) for T=tfix(s)T=t_{\text{fix}}(s).

For s1/Ns\lesssim 1/N, we find

n(s)=NU(TN)+𝒪(s2).n(s)=NU(T-N)+{\cal O}(s^{2}). (A35)

For Ns1Ns\gg 1, using Eqs. (A15) and (A19), we obtain the asymptotic expansion

n(s)NU(Tln(2Ns)+γs).n(s)\approx NU\Big{(}T-\frac{\ln(2Ns)+\gamma}{s}\Big{)}. (A36)

A.6 ξ\xi for sβ1s\beta\ll 1

From Eq. (4), using Eqs. (A27), (A35), and (A2), we obtain to first order in sβs\beta

ξ1+eNUu(s)eNU(TN)u(s)u2(s,0)sβ+𝒪(s2β2).\xi\approx 1+\frac{e^{-NUu(s)}-e^{-NU(T-N)u(s)}}{u_{2}(s,0)}s\beta+{\cal O}(s^{2}\beta^{2}). (A37)

If further NU(TN)u(s)1NU(T-N)u(s)\ll 1, we obtain

ξ1+N(1N/T)sβ+𝒪(s2β2),\xi\approx 1+N(1-N/T)s\beta+{\cal O}(s^{2}\beta^{2}), (A38)

and for TT\rightarrow\infty,

ξ1+Nsβ+𝒪(s2β2).\xi\approx 1+Ns\beta+{\cal O}(s^{2}\beta^{2}). (A39)

A.7 ξ\xi for Nsβ1Ns\beta\gg 1

For Nsβ1Ns\beta\gg 1, only the first term contributes to Eq. (2), and we obtain from Eqs. (A36) and (A3)

u2(s,β)=(1e2sβ)[1exp(NU[Tln(2Nsβ)+γsβ](1e2s))].u_{2}(s,\beta)=(1-e^{-2s\beta})\Big{[}1-\exp\Big{(}-NU\Big{[}T-\frac{\ln(2Ns\beta)+\gamma}{s\beta}\Big{]}\big{(}1-e^{-2s}\big{)}\Big{)}\Big{]}. (A40)

Likewise, in this limit we can simplify Eq. (3) to

u2(s,0)=N+1Nexp[NU(TN)(1e2s)]/Nexp[NU(1e2s)],u_{2}(s,0)=\frac{N+1}{N}-\exp[-NU(T-N)(1-e^{-2s})]/N-\exp[-NU(1-e^{-2s})], (A41)

giving

ξ(1e2sβ)[1exp(NU[Tln(2Nsβ)+γsβ](1e2s))](N+1)/Nexp[NU(TN)(1e2s)]/Nexp[NU(1e2s)].\xi\approx\frac{\displaystyle(1-e^{-2s\beta})\Big{[}1-\exp\Big{(}-NU\Big{[}T-\frac{\ln(2Ns\beta)+\gamma}{s\beta}\Big{]}\big{(}1-e^{-2s}\big{)}\Big{)}\Big{]}}{\displaystyle(N+1)/N-\exp[-NU(T-N)(1-e^{-2s})]/N-\exp[-NU(1-e^{-2s})]}. (A42)

Furthermore, for TT\rightarrow\infty, this expression simplifies to

ξ(1e2sβ)(N+1)/Nexp[NU(1e2s)].\xi\approx\frac{\displaystyle(1-e^{-2s\beta})}{\displaystyle(N+1)/N-\exp[-NU(1-e^{-2s})]}. (A43)

If NU1NU\ll 1, then ξN\xi\rightarrow N in the limit ss\rightarrow\infty.

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