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Am J Hum Genet. Aug 2003; 73(2): 336–354.
Published online Jul 11, 2003.
PMCID: PMC1182137

Finding Haplotype Block Boundaries by Using the Minimum-Description-Length Principle


We present a method for detecting haplotype blocks that simultaneously uses information about linkage-disequilibrium decay between the blocks and the diversity of haplotypes within the blocks. By use of phased single-nucleotide polymorphism data, our method partitions a chromosome into a series of adjacent, nonoverlapping blocks. The partition is made by choosing among a family of Markov models for block structure in a chromosomal region. Specifically, in the model, the occurrence of haplotypes within blocks follows a time-inhomogeneous Markov process along the chromosome, and we choose among possible partitions by using the two-stage minimum-description-length criterion. When applied to data simulated from the coalescent with recombination hotspots, our method reliably situates block boundaries at the hotspots and infrequently places block boundaries at sites with background levels of recombination. We apply three previously published block-finding methods to the same data, showing that they either are relatively insensitive to recombination hotspots or fail to discriminate between background sites of recombination and hotspots. When applied to the 5q31 data of Daly et al., our method identifies more block boundaries in agreement with those found by Daly et al. than do other methods. These results suggest that our method may be useful for designing association-based mapping studies that exploit haplotype blocks.


In the past 2 years, a series of surveys of SNPs has documented linkage disequilibrium (LD) extending across long distances in the human genome (Daly et al. 2001; Reich et al. 2001 ; Gabriel et al. 2002). This long-range LD is not confined to occasional, distantly separated marker pairs in LD; rather, these studies report a blocklike LD structure. The blocks of LD are regions, typically <100 kb, in which LD decreases very little with distance between markers. Between these blocks, however, LD is observed to decay rapidly with physical distance. Concomitant with the blocks of LD, these studies also find low haplotype diversity within blocks. Often, >90% of the chromosomes in a sample possess one of only two to five haplotypes within a block (Daly et al. 2001; Patil et al. 2001; Gabriel et al. 2002).

The existence of haplotype block structure has serious implications for association-based methods for the mapping of disease genes. On the one hand, if a causative allele occurs within a long block of LD, then it may be difficult to localize that causative gene at a fine scale by association mapping. On the other hand, if the diversity of haplotypes within blocks is low, then common disease genes may be mapped, at least to within a haplotype block, by using far fewer markers than previously imagined (Goldstein 2001). Accordingly, a variety of methods for the identification of haplotype blocks from SNP data have been proposed (Daly et al. 2001; Patil et al. 2001; Gabriel et al. 2002; Wang et al. 2002; Zhang et al. 2002b). These methods can be classified into two categories: those that divide strings of SNPs into blocks on the basis of the decay of LD across block boundaries and those that delineate blocks on the basis of some haplotype-diversity measure within the blocks.

Daly et al. (2001), Gabriel et al. (2002), and Wang et al. (2002) define haplotype blocks on the basis of LD decay between blocks. Within a 500-kb region on 5q31, Daly et al. (2001) use a hidden Markov model in which the transition probabilities between states at adjacent markers are related to the decay of LD, as measured by the statistic D. Unfortunately, their method appears to require the ability to identify a small set of “ancestral” haplotypes spanning the region of interest, so their method is not generally useful for new data sets. Wang et al. (2002) define haplotype blocks by using the four-gamete test (FGT) of Hudson and Kaplan (1985). Gabriel et al. (2002) define a block to be a region in which a small proportion of marker pairs show evidence for historical recombination. Hence, Gabriel et al.’s (2002) method is similar to Wang et al.’s (2002) method, except that the evidence for historical recombination is pairwise D<1 in the latter, whereas, in the former, the evidence for historical recombination relies on whether the upper and lower confidence limits on estimates of pairwise D fall within certain threshold values. The threshold values for these confidence limits are subjectively determined for the populations under study, making them somewhat arbitrary and not easily applied to data sets that are from different populations or that have different sample sizes.

Patil et al. (2001) and Zhang et al. (2002b) partition sequences of SNPs into blocks by using methods that are based on haplotype diversity within blocks, instead of the decay of LD across block boundaries. They locate blocks so that the most common haplotypes within blocks in a data set can be identified using the smallest number of SNPs, called, following Johnson et al. (2001), “haplotype tagging SNPs” (htSNPs). Patil et al. (2001) find block boundaries by using a greedy algorithm, whereas Zhang et al. (2002b) present a dynamic programming algorithm to find the block partition corresponding to a globally minimal number of htSNPs. These methods are appealing because they suggest that, by defining blocks in an appropriate fashion, the number of SNPs that must be typed for association studies may be reduced (Zhang et al. 2002a). Zhang et al. (2002b) also show that other measures of block quality can be used in their dynamic programming algorithm, so long as the measures of block quality are functions of the SNPs within blocks, rather than between blocks. For example, they illustrate the use of their algorithm in conjunction with Clayton’s haplotype-diversity measure (Johnson et al. 2001).

Like LD, patterns of haplotype diversity are influenced by recombination, and, therefore, block boundaries based on measures of within-block haplotype diversity should coincide with drops in LD. However, these within-block methods are not specifically designed to define blocks in terms of the drop in LD across block boundaries, and both Patil et al. (2001) and Zhang et al. (2002b) have been careful not to attach specific biological meaning to the position of block boundaries found by their methods.

Here, we present a method for the detection of blocks in phased SNP data by simultaneously using information about LD decay between blocks and about the diversity of haplotypes within blocks. We develop our method by formalizing the task of finding block boundaries as a problem in statistical-model selection, and we then apply the minimum-description-length (MDL) criterion (Rissanen 1978, 1989) to select the block designations that best capture the structure within the data. The MDL criterion is an application of information theory to statistical-model selection. The “description length” of a data set is a penalized negative log-likelihood, which is a function of the number and position of block boundaries. The best set of block boundaries, by the MDL criterion, is the set that achieves the shortest description length for the data. (An introduction to the MDL principle may be found in Hansen and Yu 2001.) Lanterman (2001) provides a discussion of the relationship between MDL and other model-selection procedures. The MDL criterion has been successfully applied to a wide variety of model-selection problems in the fields of computer science, electrical engineering, and database mining.

Computing a description length for SNP data requires a probability model for the data. Within the MDL framework, this model is intended to provide a statistical description of the data and is not necessarily designed to mimic the genetic processes generating the data (Hansen and Yu 2001). Such a descriptive framework is appropriate for inference of haplotype blocks, because, even though the etiology of the blocks is unclear, the patterns of block structure observed in empirical studies are detailed enough to be useful in forming a model. We developed a variety of different probability models for haplotype blocks in SNP data, and we report here the simplest of those models, which captures both LD decay between blocks and low haplotype diversity within blocks. In this model, the haplotypes possessed by a chromosome at adjacent blocks occur as a Markov process, with varying degrees of dependence between different types. This variable dependence is motivated by the observation of Daly et al. (2001) that the blocks in 5q31 occur at locations where there is a reduction of LD, yet considerable association between a small number of common haplotypes at adjacent blocks may still occur. The Markov chain of our model is different from that of Daly et al. (2001), in that the “time steps” in their chain were individual SNPs, whereas, in ours, the “time steps” are different blocks. Consequently, the MDL method does not rely on LD between pairs of markers; rather, it exploits LD between whole haplotypes carried at adjacent blocks. Such a block-based measure of LD should be primarily affected by recombination between blocks and less affected by multiple mutation, gene conversion, and genotyping error than are LD measures based on pairwise comparisons between loci.

In the past half-year, two other block-partitioning methods using the MDL criterion have been independently developed and presented at symposia. Although these two other methods also use the MDL criterion, they are implemented differently. Koivisto et al. (2003) report a method that uses an underlying probability model, which is different from the Markov model that we employ. In their method, haplotypes within blocks are clustered using k-means clustering, and the description length depends on the number of clusters within a block and how closely haplotypes within blocks cluster together (as well as on the number and position of blocks). Their method is thus based on a measure of within-block haplotype diversity. Unlike our method, Koivisto et al.’s (2003) model does not account for the association between common haplotypes at adjacent blocks, as observed by Daly et al. (2001). Greenspan and Geiger (2003) present an MDL-based haplotype block–partitioning method that has the attractive feature in that it accepts both phased haplotype data and unphased genotype data. When using unphased genotype data, their program is capable of estimating the underlying haplotypes. They base their description length on a Markov model similar to the one that we use, but their model underlying the distribution of types within a block is more complex, allowing for the specification of a limited number of ancestral haplotypes from which the observed haplotypes are derived by way of (possibly recurrent) mutation. This extra complexity makes minimizing the description length over the different parameters of their model a difficult task, which they tackle with a heuristic search strategy. We employ a more efficient optimization via a dynamic programming algorithm.

In the second part of the present article, we assess how our method and other methods behave when applied to simulated data. We undertake a series of simulations using the coalescent with recombination hotspots to induce LD decay at defined locations and produce a known block structure in the data. Recombination hotspots have been shown to be associated with blocks of LD in single-sperm–typing studies (Jeffreys et al. 2001; Cullen et al. 2002), and population-level data are consistent with the occurrence of recombination hotspots at block boundaries (Gabriel et al. 2002; Kauppi et al. 2003). Thus, our simulations provide an approximation to one of the possible processes giving rise to haplotype blocks. Regardless of how well these simulations correspond to the actual process that produces haplotype blocks, they do reveal useful results about the behavior of different block-finding methods. The main conclusions from these simulations are that (1) in the presence of recombination hotspots, our MDL-based method locates block boundaries at the hotspot sites and rarely places block boundaries at sites that recombine with lower intensity; (2) methods based on haplotype diversity are relatively insensitive to LD decay across block boundaries; and (3) the FGT-based method of Wang et al. (2002) locates block boundaries not only at the sites of recombination hotspots but also at numerous other, nonhotspot locations.

Finally, we apply various block-finding methods to the 5q31 data of Daly et al. (2001) and to the biallelic polymorphisms found in 86 complete human mtDNA sequences. Daly et al. (2001) originally discovered and assessed haplotype blocks in 5q31, with a significant amount of human interaction with the data. The blocks delineated by Daly et al. (2001) provide a compelling picture of block structure, which one would hope could be recovered comparably with “unsupervised” (i.e., not requiring considerable human input) block-finding methods. The blocks found by the MDL method are nearly in agreement with the original blocks found by Daly et al. (2001). The htSNP method of Zhang et al. (2002b), although largely insensitive to LD decay in the simulated data, also locates blocks in the 5q31 data, close to the blocks described by Daly et al. (2001). The FGT-based method, however, finds no fewer than 75 blocks in the 5q31 data, each with an average length of only 1.37 SNPs, suggesting that the FGT method’s utility may be reduced in the presence of the gene conversion, multiple mutation, and genotyping errors that are expected in real data. This finding is corroborated by analysis of the mitochondrial data.


In the present section, we formulate our statistical model and elaborate on the theory behind the MDL criterion. A detailed account of the computation of the description length and its optimization over possible block assignments is given in appendixes AC. The method is implemented in the program MDBlocks, available on the World Wide Web (see MDBlocks Home).

We use the MDL criterion (Rissanen 1989) to infer block boundaries from SNP-marker data typed on a sample of chromosomes. The MDL criterion relies on the relationship between the probability of a random variable and the amount of information, measured in binary digits, or “bits,” required in order to encode the values of that random variable in a binary prefix code (see Cover and Thomas 1991, chap. 5). In brief, for a random variable W taking values in a set W, a prefix code is a code that establishes an association between each value w[set membership]W and a unique code word, consisting of a series of 0s and 1s, such that no code word is the prefix of any other code word. We denote by equation M1 the length, in bits, of the code word associated with the value w. This quantity is called the “code length” of w. An efficient code—one that requires, on average, few bits in order to transmit a series of instances of W—will assign short code words to the frequent values of W and longer code words to the less frequent values of W. This is a fundamental principle of data compression. Two laws from the field of information theory, Kraft’s inequality and Shannon’s source-coding theorem, together give the result that the prefix code with the smallest expected code length for transmitting a sequence of values of W is achieved by assignment of code words to values of w, such that equation M2 for all w[set membership]W, where equation M3 is the probability mass function for W and equation M4 denotes the smallest integer equal to or larger than x. In other words, the optimal coding scheme is one that assigns a code word of length equation M5 to every w[set membership]W.

The MDL criterion exploits this connection between probability and information in the case in which the random variable in question is a data set with random features that can be described by a parameterized probability model for the data. In such a case, the code length (dispensing with the “equation M6” function and allowing for fractional bits, as we will do for the remainder of the present article) of the data set, conditional on the model and all its parameters, is the negative log-likelihood function with the log taken to base 2. Under this interpretation, the familiar maximum-likelihood estimate is identical to the value of the parameter that minimizes the code length of the observed data. The MDL framework extends more readily, however, than the maximum-likelihood framework to problems of model selection between nonnested models, because models and their parameters may also be described in terms of the code length required in order to encode them, providing a natural way of penalizing more-complex models for the addition of parameters. One way of formalizing this model-selection framework is by choosing the model that minimizes the description length for a two-stage coding of the model and the data (Rissanen 1989; Hansen and Yu 2001; Lee 2001). In a two-stage coding scheme, the description length is the sum of the code length required in order to encode the estimated values of the parameters in the model and the code length of the data set under the chosen model and parameter values.

In the next subsection (“Probability Model for Haplotypes and Blocks”), we introduce a descriptive probability model for the data. This probability model provides the basis for computing the code length of the data. The subsequent subsection (“Two-Stage Coding”) describes in more detail the two-stage coding. Appendixes A, B, and C respectively describe how we compute the description length of the parameters, how we optimize the description length over possible choices of block boundaries by using a dynamic programming algorithm, and how we treat missing data. Appendix D summarizes the notation used in the present article.

Probability Model for Haplotypes and Blocks

The data consist of N homologous chromosomes sampled from a population and typed at M SNPs each. The map location of the SNPs is used to order the SNPs from 1 to M along each chromosome, but our model does not otherwise consider the distance between SNPs. The two alleles at each SNP may be coded as 0 or 1, and we denote the allele carried on the ith chromosome at the jth SNP by equation M7, for i=1,…,N and j=1,…,M. The data set can be thought of as a matrix Y of 0s and 1s with N rows and M columns. We use the term “block” to refer to a collection of consecutive SNPs, and we describe a haplotype block structure in the data set by the specification of R adjacent, nonoverlapping blocks, indexed by k (k=1,…,R). The index of the last SNP in the kth block is denoted by equation M8, and the kth block contains equation M9 SNPs. Since the blocks are adjacent, the kth block begins at SNP index equation M10 and ends at equation M11. Thus, the right endpoints of the blocks equation M12, where equation M13, specify the block structure. Our notation follows the convention that variables superscripted by “(k)” pertain to the kth block.

We use the words “type” or “haplotype” only in reference to a particular block of SNPs: a haplotype is characterized by a sequence of 0s and 1s within a block. At the kth block, the N chromosomes in the data set will exhibit a number equation M14 of distinct haplotypes. Across all blocks, we use S to denote the vector equation M15. We denote the set of distinct haplotypes observed in the data set at block k by equation M16.

We model the occurrence of haplotypes along a chromosome, given R and E, by a simple Markov model with blocks playing the role of “time” and with time-inhomogeneous transition probabilities. In other words, the probability that a chromosome carries a certain haplotype at block k, conditional on the haplotypes carried on the chromosome at all blocks to the left of k, depends only on the haplotype carried by the chromosome at block k-1. The parameters of the Markov model are the unknown population frequencies of the types at the first block, equation M17, and the unknown population frequencies, equation M18, of haplotype pairs in adjacent blocks. The parameter equation M19 is the probability that a chromosome sampled from the population and carrying type r at block k-1 carries type s at block k. We let equation M20 be an equation M21 transition-probability matrix with entry equation M22 at the rth row and the sth column, and equation M23. The statistics of the data under the Markov model are the observed counts, equation M24, of the equation M25 different haplotypes at block 1 and, for each k=2,…,R, an equation M26 matrix equation M27 whose entry at the rth row and the sth column is equation M28, the number of chromosomes in the data set having haplotype r at block k-1 and type s at block k. Given R, E, S, P, and equation M29, the probability of the occurrence of types along every chromosome in the data set can be expressed as a function of equation M30 and equation M31. The log of the probability of the types on all the chromosomes, with the order of chromosomes within the data set considered fixed, is hence

equation image

This probability distribution does not attempt to follow the stochastic process of genealogical inheritance, mutation, and sampling in a population that gives rise to the sample. Considerable work has been done on such models of coalescence with recombination that try to capture the actual process generating the data. However, inference based on the coalescent with recombination is difficult in this particular application, both because of computational considerations and because the specific demographic or recombinational processes generating the block structure are not well known. Instead, we propose our model as a computationally tractable statistical description of the dominant features of interest in the data.

Although the pattern of largely independent haplotypes across block boundaries, with occasional association between a few types, can be reasonably modeled using a Markov chain, the above formulation demands the specification of more parameters than are typically necessary. Imagine, for example, that there is no association between haplotypes in blocks k-1 and k. It would be superfluous to specify every single equation M32 in this case, because equation M33 would be the same for all r and would simply equal the marginal frequency of the types at block k. Because the MDL criterion imposes a penalty for models with additional parameters, we consider parameterizing the Markov model with transition matrices equation M34 (as opposed to equation M35), which have their entries determined in a way that requires fewer parameters than the equation M36 required in order to specify all the elements of each equation M37. We do this by setting each value equation M38 to the marginal frequency of the sth type in the kth block, unless there is a strong enough association between one or more types across the block boundary such that the overall description length (of the data and the parameters) may be reduced by explicitly modeling that dependence.

To specify the elements of equation M39 as above, we use the unknown marginal frequencies of the equation M40 types at block k (k=2,…,R). Let the row vector equation M41 denote those unknown marginal frequencies. Also, let Δ be an equation M42 matrix of indicators equation M43, which take the value 1 if the dependence between type r in block k-1 and type s in block k is to be explicitly modeled and the value 0 otherwise. We define the transition-probability matrix equation M44 to have elements equal to those of equation M45 at all entries for which the corresponding elements in Δ are 1. For the remaining entries (corresponding to equation M46), the values of equation M47 are set to be proportional to the corresponding components of equation M48 and are scaled so that the rows of equation M49 sum to unity. That is,

equation image

where equation M50 and equation M51 respectively denote the rth rows of equation M52 and equation M53, 1 is a vector of 1s, and · is the vector dot product. This scheme allows the significant dependence between a small number of types between adjacent blocks to be modeled without having to invoke a very large number of parameters. In appendix A, we give the details of how we choose values for Δ. The log probability of equation M54 and Z, given equation M55, is given by equation (1) but with equation M56 replaced by equation M57.

The above formulas describe a probability model for the occurrence of types along a chromosome. To apply the MDL criterion, however, we must compute the code length of the whole data set, which includes the sequences of the types at each block. Thus, we need a probability model for the sequence of 0s and 1s of each type in equation M58 for k=1,…,R. Coalescent theory informs us of a reasonable form for the distribution of such sequences in a sample, but the actual calculation of that probability is computationally expensive because the number of possible coalescent topologies is so large. Some approximations have been developed (e.g., see Stephens and Donnelly 2000), but these too are difficult to implement and are also computationally demanding. Instead, we assume a simple Bernoulli model for sequences within each type in equation M59—the value of 0 or 1 at each site [ell] within a block k is drawn independently—with probability equation M60 if the site is a 1 and with probability equation M61 if it is a 0. We denote the vector of values of equation M62 for all equation M63 SNPs in block k by equation M64, and we define equation M65. When we let equation M66 denote the type of the [ell]th SNP of the sth type in the set equation M67, the code length of the list of sequences in equation M68 is computed as

equation image

Two-Stage Coding

A useful way of thinking about two-stage coding is to imagine that you wish to transmit, in binary data format, the data set of SNPs, transmitting as few bits as possible. If you naively sent only one bit (a 0 or a 1) for each value of Yi,j, then you would have to send N×M bits to transmit the whole data set. This might be a reasonable coding scheme if every SNP in the data set were a 0 or a 1, with equal probability, independently of all other SNPs; however, that is not the case. Instead, there is a structure to the pattern of 0s and 1s in haplotype data, and this structure may be exploited to minimize the total number of bits required in order to transmit the data. Such data compression may be achieved by imagining that the data set is a realization from a suitable descriptive probability model, the family of which has been agreed on ahead of time by the transmitter and the receiver. By the relationship between probability and code length, the probability model and its parameter values determine the best prefix codes to be used to denote and transmit certain features of the data set. For the transmitted message to be properly received, the values of the parameters must be sent before the rest of the data features, so that the receiver will know how to properly decode the portion of the message that includes the data set. Hence, the total number of bits required is the description length of the parameters plus the description length needed to transmit the data by using the coding scheme implied by the probability model and its previously transmitted parameter values.

We have implemented a two-stage coding scheme in which the family of descriptive models is that given by equation (1) for the blocks and that implied by equation (3) for the sequences within each block. The overall description length D1 for the first stage is the sum of the code lengths for each of the parameters, R, E, S, P*, q, and h (see appendix A). For the second stage, the types carried by each chromosome must first be transmitted by sending a series of prefix code words for each type carried at each block on each chromosome. The number of bits required in order to transmit this information is given by equation (1), with the log taken to base 2. The final part of the second stage involves transmitting the sequences of the types within each block. For all the types at all the blocks, this requires the code length given by equation (3). The overall description length of the second stage is thus

equation image

In the two-stage MDL framework, the best model is the one that minimizes the total description length, D1 + D2. Thus, there is a trade-off: increasing the complexity of the model requires more bits in order to describe the model and parameters, but, if the extra description length required for the model is more than offset by a reduction in the description length of the data themselves, then the more complex model is justified.

In appendix B, we present an algorithm for minimizing the total description length over models indexed by the number of blocks and the positions of the block boundaries. Because the method relies on successively performing a dynamic programming algorithm, we call the algorithm the “iterative dynamic programming (IDP) algorithm.” We also have developed a more computationally efficient, but approximate, algorithm that we call the “iterative approximate dynamic programming (IADP) algorithm.” The input to each algorithm is the N×M matrix of phased SNP data, and the output is the number of blocks and positions of the block boundaries corresponding to the MDL found by each algorithm. Appendix C describes how our program handles missing data by imputing the missing values in such a way as to minimize the entropy of the observed distribution of haplotypes within a block.


Coalescent Simulations with Recombination Hotspots

To evaluate our method’s ability to identify regions of rapid LD decay, we applied it to data simulated using the coalescent with recombination hotspots. Although these simulations do not recreate all of the patterns observed in haplotype data, they do create LD patterns, in the data, that are like those described in recent empirical studies. We also applied other block-finding algorithms to the same simulated data, to compare the results and behaviors of these different methods.

To simulate SNP data with recombination hotspots, we modified the program of Hudson (2002) to include variable recombination rates at sites of recombination. In the program, the scaled population recombination rate ρ for the entire simulated chromosomal segment was apportioned to equally spaced sites separating 200 chromosomal segments, and, among those sites, 10 equally spaced ones were chosen to be hotspots, having recombination rates Rh times greater than the “background” rate of recombination. We performed sets of simulations for Rh values equal to 50, 100, and 200 and ρ values equal to 200 and 500 for sample sizes of N=18 and N=100. The proportion of all recombinations that occurred within intermarker intervals containing hotspots was, on average, ~75%, ~85%, and ~90% for Rh values of 50, 100, and 200, respectively. For Rh=200, the percentage of recombinations occurring at hotspots is on the order of the value of 95%, found by Jeffreys et al. (2001) in the major histocompatibility complex (MHC) region. The presence of hotspots leads to a bimodal distribution of the number of recombinations occurring between adjacent marker pairs: most adjacent marker pairs experience few recombination events in the ancestry of the sample; however, adjacent markers separated by hotspots experience many more recombinations. Figure 1 shows an example of this bimodal distribution. In addition, we simulated five hotspots equally spaced in the first half of the chromosome and no hotspots in the second half. The conclusions regarding the behavior of block-finding methods from these simulations with 5 hotspots were qualitatively similar to those with 10 hotspots (results not shown). All simulations assumed a population of constant size.

Figure  1
Distribution of the number of recombination events between adjacent marker pairs. In 1,000 ancestries simulated from the coalescent with 10 recombination hotspots, we recorded the number of recombinations occurring in intervals between adjacent marker ...

Although the recombination locations were equally spaced, the mutations were placed randomly within the entire segment according to the infinite-sites model. Only sites with a minor-allele frequency >0.05 in a sample of 100 were included in the sample, and the number of segregating sites was chosen so that a mean of ~110 sites was obtained in each simulation replicate. Although genotyping technologies are constantly evolving, realistic marker densities are on the order of one marker every 2–5 kb, and suggested block lengths range from 5 to 100 kb (Daly et al. 2001; Patil et al. 2001; Gabriel et al. 2002). If we assume a recombination fraction of 1% per Mb and an effective size of 10,000 for the human population, then the simulations with ρ=200 are equivalent to a marker density of approximately one marker every 4.5 kb and block lengths of, on average, 45 kb. Under the same assumptions, a value of ρ=500 is approximately equivalent to one marker every 11 kb and average block lengths of 110 kb. We can also use the relationship suggested by Nordborg and Tavare (2002), wherein ρ=100 is equivalent to ~100 kb. Under this assumption, in our simulations, ρ=200 corresponds to one marker every 1.8 kb with blocks of average length 18 kb, and ρ=500 corresponds to one marker every 4.5 kb with blocks of average length 45 kb.

We applied the following four block-finding algorithms to the simulated data and compared their behavior.

Patil et al.’s (2001) htSNP-based criterion as implemented by Zhang et al. (2002b) in the program Hapblock. We used the default settings of 80% coverage and [gt-or-equal, slanted]80% of all haplotypes in a block represented more than once.
The diversity-based method suggested by Zhang et al. (2002b), using Clayton’s haplotype-diversity measure (Johnson et al. 1997). The method is implemented in Zhang et al.’s (2002b) Hapblock program, and we again used the default settings.
The method presented by Wang et al. (2002), based on the FGT. We implemented the method in the C programming language.
Our MDL-based method with the IADP algorithm, implemented in the computer package MDBlocks.

We applied all four methods to the same set of 1,000 simulated data sets with N=18. We also applied the FGT and MDB methods to 1,000 data sets of N = 100; the execution of the DB and htSNP methods was too slow to apply them to these data sets.

To describe the behavior of various methods, we calculated the proportion of all recombination hotspots that were identified as block boundaries, calling this statistic the “hotspot sensitivity.” Although none of the four methods are specifically designed to detect hotspots of recombination, the extent to which a method detects locations where LD drops sharply is relevant to SNP selection for association studies. In the simulations, LD dropped markedly at recombination hotspots, so the assessment of how well each method identifies simulated recombination hotspots is of practical interest. We also recorded the proportion of block boundaries that were not concordant with hotspots. We refer to this number as the hotspot nonconcordance (HNC) rate. Since relatively few recombinations occurred at sites that are not hotspots (fig. 1), the HNC rate measures how often methods declare block boundaries where the drop in LD is, on average, relatively small. “Sensitivity” and “nonconcordance” could be defined more loosely; for example, a block boundary could be considered concordant when it is separated from a hotspot by some small number of SNPs, instead of when it is exactly within the interval containing the hotspot. The conclusions from such a relaxed definition are similar to what is reported here for the more strict definition (results not shown).

For N=18, the hotspot sensitivity of the MDB method was comparable to that of the most sensitive methods, but the HNC rate of the MDB method was consistently lower than the HNC rates of the other methods (table 1). The differences in hotspot sensitivity between the MDB, FGT, and htSNP methods were never >15%. In contrast, the decrease in the HNC rate for the MDB method relative to the other methods is never <18% and is typically >22%. Overall, the results suggest that the MDB method performs well in the inference of block boundaries at locations with sharp drops in LD and that the MDB method also has the discriminating power to not identify as block boundaries those locations with comparatively small drops in LD. The DB method had the lowest hotspot sensitivity across all parameter sets, perhaps because our simulations did not create patterns of low haplotype diversity between hotspots. The results suggest that haplotype-diversity–based block-finding methods can be insensitive to sharp drops in LD.

Table 1
Hotspot Sensitivity and Nonconcordance for Simulated Data with N=18[Note]

HNC rates were typically highest for the FGT and htSNP methods. Both of those methods identified block boundaries at many more locations than the 10 hotspots that were simulated. For instance, with ρ=500 and Rh=200, the FGT and htSNP methods found, on average, ~20 block boundaries, whereas the MDB method found, on average, only ~10 block boundaries (table 2). This is not surprising for the FGT method, because it will declare a block boundary between any two adjacent sites with evidence of one or more recombinations in the history of the sample.

Table 2
Average Number of Blocks Inferred across Various Simulation Settings[Note]

The difference in HNC between the MDB and FGT methods is far more pronounced for N=100 (table 3). With ρ=200 and Rh=200, the mean HNC rate of the MDB method was 0.10, whereas, for the FGT method, it was 0.52. These rates reflect that, on average, the MDB method inferred 9 of the 10 hotspots as block boundaries and inferred only one additional block boundary where a hotspot did not exist. By contrast, the FGT method also typically identified as block boundaries 9 of the 10—or all 10—hotspots, but it also inferred block boundaries at an additional [gt-or-equal, slanted]10 locations that were not at recombination hotspots. Relative to the simulations with N=18, when N=100, the hotspot sensitivity is higher for both the MDB and FGT methods, and the HNC rate is lower for the MDB method, suggesting that a large sample size is valuable for the accurate location of block boundaries where sharp drops in LD occur.

Table 3
Hotspot Sensitivity and Nonconcordance for Simulated Data with N=100[Note]

A boundary declared by the MDB method is more likely to have had a high number of recombinations than a boundary declared by the FGT method or the htSNP method. This can be seen in the distribution of the number of recombinations occurring at block boundaries inferred by the various methods (fig. 2), which shows that a higher proportion of block boundaries inferred by the MDB method have had a high number of recombinations when compared to block boundaries inferred by the FGT method or the htSNP method.

Figure  2
Distribution of the number of recombinations within marker intervals identified as block boundaries, for the MDB (A), FGT (B), and htSNP (C) methods. Plus symbols (+) show the proportion of intermarker intervals, identified as block boundaries by each ...

Coalescent Simulations with Uniform Recombination

We also investigated the behavior of block-finding algorithms on data simulated with uniformly distributed recombination. The simulation parameters were identical to those in the simulations with recombination hotspots, except that Rh=1 for uniform recombination. Without hotspots, only 10%, on average, of all recombination events in the history of a sample of 100 chromosomes occurred at the 10 sites with the most recombination. The same figure for the simulations with hotspots was >75% for all values of Rh. Furthermore, in the absence of hotspots, Phillips et al. (2003) and Wang et al. (2002) have shown that the distribution of allele frequencies, marker spacing, and recombination rates is an important factor influencing the observation of block lengths similar to those observed in human data. None of these factors were manipulated in a manner to produce block structure in our uniform-recombination simulations. For these reasons, there will be less-pronounced block structure in the simulations with uniform recombination than in the simulations with recombination hotspots.

The MDB method consistently identifies block boundaries in the absence of recombination hotspots. This finding makes it clear that the MDB method is not a test for the presence of recombination hotspots. For instance, with N=100 and ρ=500, the MDB method found, on average, 18 blocks in the simulated data. That the MDB method finds blocks even when there should not be strong block structure in the data may seem to be an undesirable property of the method; however, we note that the MDB method identifies fewer blocks in the uniform-recombination case than do all three of the other methods (table 2). On average, the FGT method inferred the highest number of blocks, followed by the htSNP method, with both methods inferring at least three times the number of blocks inferred by the MDB method for almost all values of ρ and N investigated. The DB method inferred fewer blocks than the htSNP and FGT methods, but it still inferred almost twice as many blocks, on average, as did the MDB method.

Without the inclusion of recombination hotspots in the simulations, it is difficult to interpret the significance, with respect to blocks of LD, of the blocks delineated by each of the different methods. The blocks inferred by each method will, in some way, reflect the history of recombination and mutation in the sample. However, further work will be necessary to understand the relationship between the blocks inferred by any method and the history of the sample, as well as the functional significance of that relationship.

Mitochondrial and Chromosome 5 Data

To test block-finding algorithms on real data, we applied our method to a data set of 822 biallelic sites in 86 complete human mtDNA sequences (Ingman et al. 2000; Maca-Meyer et al. 2001). Because there is little evidence for widespread recombination in human mtDNA (Posada et al. 2002), it would be surprising to find block structure in mtDNA data. Accordingly, the mtDNA data serve as a useful negative control for methods that detect block structure. We applied all four block-finding methods to the mtDNA data; however, we report results only for the MDB and FGT methods. The DB and htSNP methods made slow progress in analyzing the mtDNA data set; thus, their execution was terminated, since it appeared as though they would not complete the analysis within less than several weeks. Using the IDP algorithm, the MDB method found four blocks in the mitochondrial data. The block boundaries had endpoints 279, 304, and 306. The IADP algorithm did not find any block boundaries in the data. Moreover, the description length of 19,005.5 bits for the IDP result was not very different from the description length of 19,010.9 bits for the IADP result of no block boundaries. We can infer from the small difference in description length that the few blocks found by the IDP algorithm are not particularly strong block boundaries.

In contrast, the FGT method inferred the presence of 91 blocks in the mtDNA data. The inferred blocks likely occur where pairs of sites carry all four possible haplotypes owing to multiple mutations or genotyping error. This appears to be a limitation of using such a stringent definition for blocks as the FGT applied in a pairwise manner to SNPs. The contrast in the number of boundaries called by the MDB and FGT methods is still present when we exclude from the data set segregating sites in the D-loop (results not shown).

To analyze human data with known block structure, we applied the MDB, FGT, DB, and htSNP methods to the data of Daly et al. (2001). The data consist of 103 SNP markers from 5q31, haplotyped on 516 chromosomes from case and control trios in an association study of Crohn disease. The data set includes many missing values. Of the 103×516=53,148 SNPs typed in the sample, 12.7% are missing data, and 7.3% could not be unambiguously haplotyped. For the MDB method, we handled the missing data by using a heuristic approach of filling in missing data points in proportion to their associations among known subsequences of length 10 (i.e., w=10; see appendix C). Daly et al. established the existence of blocks in the 5q31 data by a variety of criteria and techniques. They first defined blocks on the basis of haplotype diversity by using an ad hoc criterion involving the ratio of observed to expected heterozygosity. They then confirmed that their block boundaries correspond to regions of high historical recombination by applying a hidden Markov model under the assumption that four classes of ancestral haplotypes across the whole region are known. Finally, they noted that jumps in the value of multiallelic D (calculated in a sliding window, over SNPs) occur at locations that they identified as block boundaries. Overall, the blocks designated by Daly et al. may be taken as a characterization of well-established block structure. Daly et al. reported the following set of block boundaries for their data set: {8–9,14–15,24,35,40,45,76–77,84–85,91,98}, where dashes indicate block-region boundaries that are imprecise because Daly et al. occasionally allowed single markers to not be included in any block.

We analyzed the 5q31 data six different times by the MDB method using the IDP algorithm. Each time, different random-number seeds were used to impute the missing data (appendix C). Across these six different imputations of the missing data, the block boundaries found by the MDB method were unchanged except for the movement of one boundary position. The resulting block boundaries were {8,14,24,36,47,57,76,86,91,98}, with 57 being replaced by 61 in two of the six runs. The average description length over the six runs was 11,556 bits.

The MDB method using the IADP algorithm produced results similar to the IDP algorithm, but in much less running time. When the IADP algorithm was used, all of the boundaries were identical to those identified by the IDP algorithm, except that the boundary at 57 was never moved to 61, and the average description length was 11,579 bits. These block designations compare favorably to those of Daly et al. (2001). Both the block calls made by Daly et al. and those made by the MDB method recognize the common boundaries {8,14,24,35/36,45/47,76,85/86,91,98}, where slash marks denote corresponding but not identical boundaries. The three cases (35/36, 45/47, and 85/86) for which the MDB method calls a boundary that is close to but not exactly concordant with one of Daly et al.’s block boundaries correspond to the three boundaries with the lowest historical recombination rates estimated by Daly et al. Furthermore, as shown in their figure 2, there is little support in the data for a distinction between a boundary at 45 versus 47 and, likewise, for a boundary at 85 versus 86. Specifically, transitions between Daly et al.’s ancestral haplotype states observed at high frequency ( >2% of chromosomes) occurring at boundary 45 would be indistinguishable from transitions occurring at boundary 47; transitions would be similarly indistinguishable between boundaries 85 and 86. The only boundary not found by the MDB method but found by Daly et al. was 40. The only boundary found by the MDB method but not by Daly et al.’s method was the boundary identified as either 57 or 61. Although the Daly et al. and MDB boundaries are generally in agreement, the description length obtained using the Daly et al. block boundaries is >850 bits longer than that obtained using the MDB method (see table 4). The large difference in the description lengths indicates that the small differences between the two sets of block boundaries are in some sense significant; however, there is no formally established method of relating description-length differences to statistical certainty.

Table 4
Results of Application of Each Method to the Daly et al. Data

The DB method found five blocks in the 5q31 data. The block boundaries were {24,44,77,88}. The first three boundaries correspond closely with boundaries identified by Daly et al. (2001).

The htSNP method found 10 blocks in the 5q31 data. The block boundaries were {8,9,11,24,40,63,80,82,91}. Most of these boundaries correspond with Daly et al.’s (2001) block boundaries; boundaries in the set {11,63,80} do not correspond. Relative to its behavior in simulations, the htSNP method appears to perform better with real data. This difference in performance may be because the simulations do not re-create observed patterns of within-block diversity.

The FGT method found 75 blocks in the Daly et al. data set. The numerous incompatibilities in this data set cause the FGT to declare a large number of block boundaries—so many, in fact, that, on the basis of these results, one might be led to conclude that there are no long blocks in the Daly et al. data. The FGT method likely performs poorly on the 5q31 data because, like the mtDNA data set, the Daly et al. data set probably has sites with multiple mutations and with genotyping errors that create incompatibilities that are not the result of recombination. In addition, nuclear sequences experience gene conversion that can cause additional incompatibilities that are not the result of reciprocal recombination.

Running on a 1.66-GHz AMD Athlon processor, to find blocks in the 5q31 data, the MDB method required 2 min 45 s when the IADP algorithm was used and 45 min when the full IDP algorithm was used. For comparison, the FGT method required <1 s, the DB method required 31 min, and the htSNP method required 65 min. In the analysis of the mtDNA data set, the IADP algorithm ran overnight, whereas the IDP algorithm took ~1 wk. Since the IADP algorithm is much faster than the IDP algorithm, performs well on simulated data, and yields a similar result to the IDP algorithm in most cases, we recommend the IADP algorithm when computational efficiency is an issue.


We have proposed a haplotype block–detection method with many advantages over existing methods. We have formulated our method as a problem in statistical-model selection by use of the MDL principle. This allows us to invoke a probability model that captures both of the empirically described, dominant features of haplotype blocks: the decay of LD across block boundaries and the low haplotype diversity often encountered within blocks. The method, which we have implemented in the computer package MDBlocks, does not depend on population-specific or arbitrarily set thresholds. Furthermore, in factoring LD into its block-boundary designations, our MDL-based method incorporates information from numerous sites within a haplotype block at once and, in doing so, avoids being overly sensitive to the patterns of variation observed in LD measures based on pairwise comparisons. If the decay of LD across block boundaries is an important determinant of the degree to which blocks are useful in gene mapping, then our MDL-based method might be preferable to methods based solely on within-block haplotype diversity.

Using computer simulations, we show that MDBlocks reliably locates the boundaries between blocks at regions of rapid LD decay. We generated data by coalescent simulations with recombination hotspots across a range of recombination rates and hotspot intensities. The method detected most hotspots, particularly when the sample size was large (100 individuals) and the intensity of the hotspots was high. Simulations show that the method discriminates between the hotspots and regions having lower, background rates of recombination. This is important because empirical studies in the MHC region (Jeffreys et al. 2001; Cullen et al. 2002) have noted that, even though recombination hotspots will strongly structure the LD in a region, recombination does not occur exclusively at hotspots. In the absence of recombination hotspots in our simulations, MDBlocks continues to find blocks, demonstrating that it does not provide a test for the existence of recombination hotspots. However, MDBlocks identifies fewer blocks than other block-finding methods in data simulated with uniform recombination.

Application of MDBlocks to two published data sets confirms that it effectively identifies haplotype block structure when it clearly exists. The MDL-based method found a set of block boundaries that were nearly in complete agreement with those identified by Daly et al. (2001) for their data on 5q31. In mtDNA data, MDBlocks finds few blocks, as one would hope, considering that there is little evidence for recombination in the mitochondrial genome.

We applied three other block-finding algorithms to the same simulated and real data, concluding that our method captures the underlying block structure more effectively than any of these methods. The FGT method of Wang et al. (2002), based on the FGT, divides both the simulated data and the real data into many small blocks. Block boundaries determined by the FGT method do not necessarily correspond to locations with large drops in LD. The consequences of such undiscerning block-finding behavior are clear when the FGT method is applied to the 5q31 data of Daly et al. (2001); despite clear evidence for the existence of ~10–12 extended haplotype blocks in that data set, the FGT method finds 75 blocks, each one containing, on average, fewer than two markers. Application of the FGT method to mtDNA data further demonstrates that the stringent pairwise nature of the FGT method makes it particularly susceptible to mistaking as block boundaries the results of genetic and sampling processes that may have little to do with recombination.

The two haplotype-diversity–based block-finding methods implemented by Zhang et al. (2002b) are largely insensitive to recombination hotspots in the simulated data. Across all parameter sets in our simulations with recombination hotspots, the block boundaries declared by these two methods (htSNP and DB) were among the least likely to be at recombination hotspots. We emphasize that these methods based on within-block diversity are not specifically designed to locate block boundaries where LD decays. Nonetheless, it is interesting and important that even the LD breakdown simulated under the assumption of recombination hotspots does not induce block structure that these methods are capable of detecting in the data.

Although the performance of our MDL-based method on simulated and real data is encouraging, we suggest here several areas of research that may lead to better methods for the detection of haplotype blocks. First, like the htSNP and DB methods, MDBlocks finds a single set of block boundaries but does not provide a summary of other sets of block boundaries that fit the data well. The description length itself, of course, provides a measure by which to compare different sets of boundaries, but determination of CIs on the basis of description lengths is an area that awaits further theoretical research. Sampling-based approaches may be helpful, in the future, for assessing the degree of confidence that may be placed in any particular set of blocks. Second, as currently implemented, our MDL method is not invariant to the direction in which SNPs are read along the chromosome. It is possible to obtain a slightly different set of blocks from MDBlocks, for any particular data set, when the data are read from left to right, rather than from right to left, along each chromosome. It might be possible, by imposition of conditions on the form of the matrix P, to ensure that the same result is achieved regardless of the direction the data are read, and this remains an unsolved problem. In the meantime, even though it is inelegant to have a description length that depends on the direction in which the markers are read, we have demonstrated that the method performs well regardless of the direction in which the data are read. Third, our method, like the FGT, htSNP, and DB methods, does not use physical distance between markers in the characterization of haplotype blocks. This is partly because the occurrence of hotspots makes the map between physical and genetic distance highly variable at fine scales (Jeffreys et al. 2001), so physical distance may not be a reliable criterion for block structure. Finally, our method, as well as the FGT, htSNP, and DB methods, requires that the phase of the markers be known, so some method of haplotype reconstruction must be applied to the data before applying these methods to it. Including the inference of haplotype blocks or the inference of recombination hotspots into the actual machinery for haplotype inference could provide some advantages, as noted by Niu et al. (2002) and as demonstrated by Greenspan and Geiger (2003).

We have presented an MDL-based method with desirable properties relative to such methods as those based on the FGT or on minimization of the number of htSNPs. Our simulations suggest that the MDL criterion, in conjunction with an appropriate statistical model, can be used to discern blocklike LD structure in the genome. Further work will be required in order to assess the degree to which the quality of the block inference is affected by the underlying statistical model and, specifically, how the performance of our method compares with that of the other recently developed MDL-based approaches of Koivisto et al. (2003) and Greenspan and Geiger (2003). A fruitful area of research may be found in the combination of elements of the model and optimization methods used here with the other MDL-based methods. In particular, Koivisto et al.’s (2003) method might be modified to include Markov dependence between a small number of their haplotype clusters defined in adjacent blocks. Additionally, the IADP algorithm with equation M69 time complexity (eq. [B2], in appendix B) may provide a means to improve the optimization search strategy of Greenspan and Geiger (2003).

Although further development is possible, it is already clear from the results of the present study that the MDL approach is well suited to the problem of identifying haplotype blocks. In general, block-finding methods will be useful in exploring the population- and genome-level processes that give rise to observations of haplotype block structure. In particular, these methods may offer a better understanding of the relative roles that recombination-rate heterogeneity and population history play in shaping the patterns of LD in human populations. Finally, descriptions of haplotype blocks may be useful in improving the feasibility of large-scale gene-mapping studies and in aiding to their design.


We thank Montgomery Slatkin, for discussions that led to the identification of the problem; Terry Speed, for suggesting the MDL approach and offering valuable advice; and Bin Yu and Michael Stumpf, for helpful discussions. We also gratefully acknowledge Mikko Koivisto, Gideon Greenspan, and two anonymous reviewers, for comments on our manuscript. This research was supported by National Institutes of Health grant GM-40282 (to M. Slatkin) and by a Howard Hughes Medical Institute Predoctoral Fellowship (to J.N.).

Appendix A: Code Lengths of the Parameters

By summing the code lengths for each parameter, we obtain the description length D1 for the first stage in our two-stage coding scheme:

equation image

To compute the code length for a parameter, one must assume a prior distribution for it. This has the form of a Bayesian prior distribution but need not have a Bayesian interpretation (Lanterman 2001; Lee 2001). In most cases, we apply Laplace’s principle of indifference and assume that the unknown parameters are uniformly distributed throughout their range. Under the assumption that R is equally likely to take any value between 1 and M, it requires equation M70 bits to encode R. Given R, there are equation M71 different values that the vector E may take. Under the assumption of a uniform distribution on E,

equation image

Since equation M72 may take any value between 1 and N, we have equation M73. Finally, each equation M74 is a real number, but, given the equation M75 types in equation M76, it would be possible to estimate it only to the precision of equation M77. Therefore, we calculate the code length for specification of equation M78 only to that precision. Following the method of Hansen and Yu (2001), equation M79. We assume that the code lengths for equation M80 are additive within and across blocks, so that equation M81 is the sum of equation M82 over all k and [ell].

The equation M83 value that minimizes the code length of the data is the maximum-likelihood estimate of the frequencies of the equation M84 types. This is a function of equation M85, the vector of observed counts of different types, taken to be in decreasing order (i.e., equation M86). Therefore, the code length for equation M87 depends on equation M88 and a prior distribution for equation M89. We consider three different prior distributions that give the following three different code lengths:

equation image
equation image


equation image

where Cj is the number of types represented by j sequences in the sample and equation M90 denotes a Stirling number of the third kind (Johnson et al. 1997). The code length equation M91 arises by assuming that each component of equation M92 is uniformly distributed, conditional on the value of the preceding component; it favors haplotype frequencies in which a small number of types account for most of the chromosomes in the sample. The code length equation M93 arises by assuming that every configuration of chromosomes into the equation M94 types is equally probable; it favors haplotype frequencies in which all the types within a block are at comparable frequencies. The code length equation M95 arises by assuming the multivariate Ewens distribution as the prior distribution for haplotype frequencies (i.e., by assuming the distribution of types within nonrecombining segments expected under neutral evolution). We choose the code length that best matches the data for each block k. To do so, we specify an indicator equation M96 that we set to 1, 2, or 3, according to which prior distribution yields the shortest code length for equation M97. Specification of equation M98 requires an additional equation M99 bits.

To encode the rth row of equation M100, we must first specify the number of 1s in the row. We denote this quantity by equation M101. Often, there will be no 1s in the row, so we specify equation M102 by first committing a single bit to indicate whether equation M103 or equation M104. Then, in the latter case, we require another equation M105 bits in order to specify equation M106; therefore,

equation image

For equation M107, there are equation M108 ways of arranging the equation M109 1s into the equation M110 entries of the rth row of equation M111. If each of those arrangements is assumed to have equal prior probability, then

equation image

bits. The total code length for equation M112 is a sum over rows: equation M113.

Determining equation M114 and Computing equation M115

Although it is easy to compute equation M116 for any equation M117, we must still find equation M118 for each block, k=2,…,R, that minimizes the description length associated with the kth block:

equation image

where equation M119 is the code length required in order to specify the values of equation M120 for which the corresponding elements in equation M121 are equal to 1, equation M122 are elements of equation M123 whose values are determined by equation M124 and equation M125 as shown in equation (2), and G is a sum of terms that are constant with respect to equation M126 and hence do not factor into the optimization over possible values of equation M127. Minimization of equation (A3) can be done by minimizing over equation M128, for each r, the quantity

equation image

where, as in the main text, equation M129 and equation M130 denote the rth rows of the matrices equation M131 and equation M132, respectively, and equation M133 is the code length for those entries in the rth row of equation M134 for which the corresponding elements in equation M135 equal 1. The number of possible configurations of 1s and 0s in equation M136 is equation M137, making the problem of determining the optimal equation M138 exponential in equation M139. As an alternative, we present a linear-time approximate algorithm for finding equation M140. In this algorithm, we order the haplotypes at the kth block in decreasing order of the distance between the observed conditional distribution of types given the haplotype in the previous block and the marginal distribution of types. We then cycle over the haplotypes in that order, and we set the corresponding values in equation M141 to 1 if doing so decreases the overall description length for the data and model parameters at the kth block.

The approximate algorithm is defined more formally after the establishment of the following notation. As before, equation M142 is a vector where equation M143 is the observed proportion of chromosomes having haplotype r in block k-1 and haplotype s in block k. We consider an ordering ν of the equation M144 types in block k in decreasing order of their contribution to the cross-entropy between equation M145 and equation M146. The newly ordered vector is equation M147, and equation M148. Let equation M149 be the vector equation M150, in which the first [ell] components may be 0 or 1 and the remaining components, equation M151, are all 0. The code length of equation M152 is

equation image

where n[ell] is the number of 1s in equation M153. This expression is consistent with equation (A2), because equation M154, as it should. Also, denote by equation M155 the values of equation M156 for which the corresponding entries in equation M157 are 1. If equation M158, then equation M159 will be a maximum-likelihood estimate given by equation M160, where equation M161 is the total number of chromosomes having type r at block k-1; thus, equation M162, and

equation image

Finally, the code length of the haplotypes at the kth block on all chromosomes carrying type r at block k-1 is

equation image

where the values for equation M163 are obtained by application of equation (2), but with equation M164 in place of equation M165. Three more notational conventions are useful: given the vector equation M166, we define equation M167 to be equation M168, with equation M169 set to 1 instead of 0; equation M170 denotes the values of equation M171 for which the corresponding entries in equation M172 are 1; and equation M173 extends equation M174 in the obvious way.

With the above definitions established, the algorithm may be described as follows.

  • Initialize: equation M175; equation M176: equation M177; and n0←0.
  • Cycling over [ell] from 1 to equation M178: if equation M179 < equation M180, then equation M181, and n[ell]n[ell]-1+1; otherwise equation M182, and n[ell]n[ell]-1.
  • End: equation M183, and equation M184.

The value equation M185, found by the algorithm, is then used as equation M186 to compute equation M187. Additionally, equation M188 is computed as equation M189, using equation (A4). The above algorithm is applied to each equation M190, and the overall code lengths used in the calculation of D1 in equation (A1) are equation M191 and equation M192.

Appendix B: Optimization

We seek to minimize D=D1+D2 over all possible choices of R and E. To do so, it helps to express D as a function of R and E. The form of the resulting function is

equation image


equation image
equation image


equation image

The last line above follows on the basis that equation M193 is a function of R. Minimization of equation (B1) with respect to R and E requires a different approach than that of Zhang et al. (2002b), because our model involves a sum over functions that are dependent on two blocks at a time. The Zhang et al. (2002b) solution is for minimization of objective functions with a simpler form. Because equation M194 is not a linear function of R, it is not possible to efficiently minimize equation (B1), but, if we assume a value of equation M195—for example, equation M196—then we can specify a dynamic programming algorithm to globally minimize the latter two terms of equation (B1). The algorithm can be described in terms of a dynamic programming matrix D with the elements assigned as follows:

equation image

The minimal value of equation M197 is equation M198, and the values of R and E minimizing the description length can be obtained by a trace-back algorithm. The optimality of the algorithm for minimization of equation M199 with equation M200 fixed at equation M201 can be proved by induction. The time complexity of this algorithm is equation M202, and the space complexity is equation M203. Generally, equation M204 is small relative to the value of equation M205, so the assumed value has little effect on the solution; nonetheless, to explore a reasonable set of assumptions for RA, we take an iterative approach. We first minimize equation M206 with RA set to 1, calling the number of blocks found equation M207. We then set RA equal to equation M208 and repeat until RA converges to a single value. Very rarely, the value of RA begins cycling, and then we choose, from the cycle, the solution that has the lowest description length. We refer to the combination of a dynamic programming algorithm with iteration over the values of RA as the “IDP algorithm.”

To improve computation speed, we have also developed an approximation to the dynamic programming algorithm that has time complexity equation M209 and space complexity equation M210. This approximate algorithm, with equation M211 fixed at equation M212, is defined recursively as

equation image

for 1[less-than-or-eq, slant]j<M. The resulting approximation to the MDL is GM, and the minimizing values of R and E are obtained by a trace back, through the assignments, to vector G. The algorithm will work well to the extent that GLEj is the same as the minimizing second-to-last endpoint (i.e., equation M213 for 0<j<k[less-than-or-eq, slant]M). In the majority of simulations, the performance of the approximate algorithm is similar to that of the full dynamic programming algorithm. As above, we iterate this approximate algorithm over values of equation M214, and we refer to this algorithm as the “IADP algorithm.”

Appendix C: Missing Data

It is not typical to obtain perfect haplotype data from autosomal regions, so most data sets will have a number of holes in them. Little work has been done on the formal treatment of missing data from within the MDL framework. To apply our method with missing data, we employ a simple, iterative method for the imputation of missing values according to their frequency of occurrence in the rest of the data set, given a window of neighboring sites. These imputed values are used to generate a set of candidate sequences at each block, and then haplotypes with data missing in each block k are assigned to these candidate sequences in a manner such that the entropy of the distribution of types at block k in the data set is minimized. First, the imputation method is described below, and then the method for assigning haplotypes to the candidate sequences is described.

Imagine that the ith chromosome is missing data at the jth SNP. We fill in the missing data at Yi,j on the basis of a “window” of size w, as follows:


Identify all sets of w SNPs not missing in chromosome i such that each member of the set is separated from the jth SNP by no more than w-1 sites not missing in chromosome i. There will be, at most, w+1 such sets (in cases in which j is near one of the ends of the string of SNPs in the data set there could be fewer than w+1). Let the indexes of the SNPs in such a set be denoted J+. These are like “windows” of nonmissing sites around j.


Let K0 be the set of all chromosomes having nonmissing data of allelic value 0, at site j, and for which no SNP at a site in J+ carries an allele different from that on chromosome i. Let K1 be defined similarly for chromosomes with the 1 allele at position j. For any chromosome k, let pk denote the number of nonmissing sites it has in J+, and define P0 as equation M215 and P1 as equation M216.


For each of the w+1 (or fewer) sets J+, compute the pair equation M217 and equation M218, and denote by equation M219 the pair of those quantities that maximizes equation M220 over the different sets J+.


Fill in the hole Yi,j with a 0 for probability Q*0 and a 1 for probability Q*1. On filling in that hole, consider it, for the remainder of the time, as nonmissing data.

Holes in the data are filled in by the above routine in random order. Some of the missing data are due to unresolved heterozygosity at a SNP. In such a case, a procedure like the one above is used, but the pair of two chromosomes from the same individual are considered simultaneously, with one of them receiving a 0 and the other receiving a 1.

We do not use the data set with missing data imputed directly. Rather, to minimize the stochastic effects of imputing the missing data, we add another step in dealing with the missing data. This occurs at the stage when the number of different types in a block is being calculated and proceeds as follows: First, types in a block are defined as if the imputed data were real. Each of these types is defined as a complete sequence of 0s and 1s. There are equation M221 such sequences at the kth block. A chromosome in the sample is said to be “compatible” (borrowing the terminology from Zhang et al. 2002b) with such a sequence if it matches that sequence at all sites that are not missing data on the chromosome. Of the equation M222 sequences, one of them will be compatible with more chromosomes than any other; all chromosomes compatible with that maximally compatible sequence are assigned the type of that sequence. Of the remaining T-1 sequences, one is compatible with the largest number of remaining chromosomes; all chromosomes compatible with it are assigned its sequence, and so forth, until there are no remaining chromosomes to be assigned. The number of sequences to which chromosomes were assigned is equation M223, which is typically less than equation M224. Assignment of types in this way is a type of greedy algorithm for assignment of blocks with missing data to compatible types in a way that minimizes the entropy of the observed distribution of types in a block.

Appendix D: Notation for Variables, Parameters, and Functions

equation M225:
Set of unique sequences observed in the data at the kth block.
equation M226:
The ceiling function. equation M227 is the least integer greater than or equal to x.
Description lengths of the first stage, the second stage, and both stages, respectively. D=D1+D2.
equation M228:
equation M229 matrix of 0s and 1s. The value of each element equation M230 dictates how equation M231 is determined by equation (2).
equation M232:
Vector equation M233, which is the rth row of equation M234.
Vector equation M235 of right endpoints of the blocks. Note that equation M236.
Collection of vectors equation M237. Each vector is of the form equation M238, where equation M239 is the proportion of 1s among the sequences in equation M240 at the [ell]th SNP in the kth block.
equation M241:
Indicator of which of the three distributions is used to calculate equation M242.
equation M243:
Number of SNPs included in the kth block. equation M244.
Number of SNPs typed on each chromosome.
Number of chromosomes typed in the sample.
equation M245:
Number of 1s in the rth row of equation M246.
equation M247:
The code-length function.
equation M248:
equation M249 matrix of elements equation M250. Each equation M251 is the unknown population frequency of chromosomes having type r in block k-1 and type s in block k.
equation M252:
equation M253 matrix of elements equation M254. The elements of this matrix are determined by equation M255, equation M256, and equation M257, as shown in equation (2), and are used in equation (4) to compute D2.
equation M258:
Vector equation M259, which is the rth row of equation M260.
equation M261:
Vector equation M262 of unknown marginal frequencies in the sampled population of the equation M263 types at block k.
Number of blocks.
Vector equation M264. Each equation M265 is the number of distinct haplotypes observed in the data at block k.
equation M266:
Vector equation M267 counting the number of times each of the equation M268 types is observed in the data at the kth block.
N×M matrix of observed SNP values. Yi,j is 0 or 1 according to the type of the jth SNP in the ith individual.
equation M269:
equation M270 matrix of observed pairs of types at adjacent blocks. equation M271 counts the number of chromosomes in the sample having type r at block k-1 and type s at block k.

Electronic-Database Information

The URL for data presented herein is as follows:

MDBlocks Home, http://ib.berkeley.edu/labs/slatkin/eriq/software/mdb_web/index.htm(for the MDBlocks program, as well as for information on the preparation of the 5q31 data of Daly et al. and the mtDNA data used in the analyses presented here)


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