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Investigating the performance of AIC in selecting phylogenetic models

Dwueng-Chwuan Jhwueng , Snehalata Huzurbazar , Brian C. O’Meara and Liang Liu EMAIL logo

Abstract

The popular likelihood-based model selection criterion, Akaike’s Information Criterion (AIC), is a breakthrough mathematical result derived from information theory. AIC is an approximation to Kullback-Leibler (KL) divergence with the derivation relying on the assumption that the likelihood function has finite second derivatives. However, for phylogenetic estimation, given that tree space is discrete with respect to tree topology, the assumption of a continuous likelihood function with finite second derivatives is violated. In this paper, we investigate the relationship between the expected log likelihood of a candidate model, and the expected KL divergence in the context of phylogenetic tree estimation. We find that given the tree topology, AIC is an unbiased estimator of the expected KL divergence. However, when the tree topology is unknown, AIC tends to underestimate the expected KL divergence for phylogenetic models. Simulation results suggest that the degree of underestimation varies across phylogenetic models so that even for large sample sizes, the bias of AIC can result in selecting a wrong model. As the choice of phylogenetic models is essential for statistical phylogenetic inference, it is important to improve the accuracy of model selection criteria in the context of phylogenetics.


Corresponding author: Liang Liu, Department of Statistics and Institute of Bioinformatics, University of Georgia, 101 Cedar Street, Athens, GA 30606 USA, Phone: +1-706-542-3309, Fax: +1-706-542-3391, e-mail:

Acknowledgments

We thank David Posada for the helpful discussion on phylogenetic model selection. We thank Diego Darriba for his generous help with implementing the phylogenetic model selection program jModelTest. Jhwueng’s research was supported by the National Science Council Award #NSC-101-2118-M-035-001 Taiwan and Postdoctoral Fellowship at the National Institute for Mathematical and Biological Synthesis (NIMBioS), an Institute sponsored by the National Science Foundation, the U.S. Department of Homeland Security, and the U.S. Department of Agriculture through NSF Awards #EF-0832858 and #DBI-1300426, with additional support from The University of Tennessee, Knoxville. We also thank NIMBioS Working Group-Gene Tree/Species Tree Reconciliation for providing the opportunity that the authors of this paper could meet to discuss this project. This research was partially supported by the National Science Foundation (DMS-1222745 to LL). Huzurbazar’s research was supported by a grant to the University of Wyoming from the National Science Foundation under grant DMS-1100615. Huzurbazar’s contribution was based upon work partially supported by the National Science Foundation under Grant DMS-1127914 to the Statistical and Applied Mathematical Sciences Institute. Any opinions, findings, and conclusions or recommendations expressed in this material are those of the author(s) and do not necessarily reflect the views of the National Science Foundation.

Appendix A

A1 The likelihood function of general substitution models

Given the DNA sequence data D of size N×M where each element of D, Duv represents the vth (v=1, …, M) nucleotide of sequence u (u=1, …, N). There are 4N possible nucleotide patterns for each column in data D. Let pi be the probability of observing the ith nucleotide pattern. The probability pi is the summation of the product of transition probabilities {P(t)=Pij(t); i, j=A, C, G, T} (see Figure A1). Thus, pi=pi(ϕ) is a function of model parameters ϕ=(θ, τ, b) where θ represents the parameters in the substitution model (typically the rate matrix Q, and the equilibrium vector π=(πA, πC, πG, πT)). When substitution rates are variable over sites, the heterogeneity of rates such as invariant parameter I, and the gamma rate parameter γ will be included in the substitution models. The tree topology τ and its branch lengths b are also treated as parameters for phylogenetic tree estimation. The rate matrix, Q, describes the rate at which bases of one type change into bases of another type. The rate matrix has the following structure

Figure A1 The probability density function p(·) of a column of nucleotides given the phylogenetic tree. (A) The column of nucleotides (A, A, C, T) are at the tips of the tree. The probability of (A, A, C, T) given the tree (on the top) is equal to the sum of the probabilities at the bottom, in which the nucleotides at the internal nodes of the tree are given. Each probability at the bottom is the product of Pij(t)s, the probabilities for individual branches on the tree. (B) A four taxa rooted phylogenetic tree. The ith site observed at tip has pattern AACT. The internal nodes zk∈{A, C, G, T}, 5≤k≤7 are ancestral status; and bj≥0, 1≤j≤6 are branches lengths.
Figure A1

The probability density function p(·) of a column of nucleotides given the phylogenetic tree. (A) The column of nucleotides (A, A, C, T) are at the tips of the tree. The probability of (A, A, C, T) given the tree (on the top) is equal to the sum of the probabilities at the bottom, in which the nucleotides at the internal nodes of the tree are given. Each probability at the bottom is the product of Pij(t)s, the probabilities for individual branches on the tree. (B) A four taxa rooted phylogenetic tree. The ith site observed at tip has pattern AACT. The internal nodes zk∈{A, C, G, T}, 5≤k≤7 are ancestral status; and bj≥0, 1≤j≤6 are branches lengths.

    A     B   G   TQ=ACGT(μAμCAμGAμTAμACμCμGCμTCμAGμCGμGμTGμATμCTμGTμT)

where μxy represents the transition rate from base x to base y and the diagonals of the matrix are chosen so that the rows sum to zero: μx=–Σ{y|y≠x}μxy. The equilibrium row vector π must satisfy πQ=0. Let P(t) be the transition probability matrix, in which Pxy(t) is the probability of base x changing to y after a period of time t. Since this is a continuous time Markov Chain, the transition probability matrix satisfies a first order ordinary system differential equation P′(t)=QP(t), to which the solution is P(tQ)=eQt. The substitution models considered in this paper are time reversible. Thus, the rate matrix Q can be diagonalized and has real eigenvalues. It is straightforward that all elements Pxy(t) in the transition probability matrix P(t) have a continuous second order partial derivative with respect to t and parameters in the rate matrix Q.

As an example, we express probability pi for a particular pattern {AACT} observed at the tips of a 4-taxon rooted tree (Figure A1B). Let zk∈{A, C, G, T}, 5≤k≤7 be the ancestor status, let bj>0,1≤j≤6 be the branch lengths. Let π be the equilibrium base frequencies, and we assume that the nucleotides at the root of the tree have reached the equilibrium frequencies π. The probability of observing pattern AACT at the tips of the tree is a sum over all possible assignments of nucleotides to internal nodes, i.e.,

(A1)pi=Pr[{AACT}|τ,b,Q]=z7z6z5{πz7Pz7z5(b5|Q)Pz7z6(b6|Q)Pz5A(b1|Q)Pz5A(b2|Q)Pz6C(b3|Q)Pz6T(b4|Q)}, (A1)

where Pxy(bjQ) is the transition probability that nucleotide y is substituted for x over a branch length bj. Equation (A1) has 43=64 terms. In general the expression for k species will have 22k2 terms. As pi is the sum of multiplications of equilibrium frequencies π and transition probabilities Pxy(b), pi has a continuous second-order partial derivative with respect to branch lengths b and parameters in the rate matrix Q and equilibrium frequency vector π.

Let ξ={ξ1,ξ2,,ξ4N} be the frequencies of 4N patterns observed in data D. Assuming that the sites evolve independently along the lineages of a phylogenetic tree, the probability density function of data D in terms of the frequencies ξ of 4N nucleotide patterns can be expressed as

(A2)f(D|ϕ)=i=14N(pi)ξi. (A2)

It follows from (2) that the log likelihood function L is

(A3)L(ϕ|D)=i=14Nξilnpi. (A3)

The log likelihood function for a fixed model m∈Ω is denoted by Lm(ϕmD), or simply by Lm. If the tree topology τ is given, pi has continuous second-order partial derivatives with respect to parameters ϕ. It follows immediately that the likelihood function Lm(ϕmD) has continuous second-order partial derivatives with respect to ϕ as Lm is the sum of the weighted logarithm of pi shown in (A3).

A2 The flow chart for estimating the expected KL divergence

Figure A2 The flow chart for estimating the expected KL divergence.
Figure A2

The flow chart for estimating the expected KL divergence.

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Published Online: 2014-5-27
Published in Print: 2014-8-1

© 2014 by De Gruyter

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