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TAG. Theoretical and Applied Genetics. Theoretische Und Angewandte Genetik
Theor Appl Genet. Aug 2009; 119(3): 397–416.
Published online May 13, 2009. doi:  10.1007/s00122-009-1047-9
PMCID: PMC2715462

Homoplasy corrected estimation of genetic similarity from AFLP bands, and the effect of the number of bands on the precision of estimation

Abstract

AFLP is a DNA fingerprinting technique, resulting in binary band presence–absence patterns, called profiles, with known or unknown band positions. We model AFLP as a sampling procedure of fragments, with lengths sampled from a distribution. Bands represent fragments of specific lengths. We focus on estimation of pairwise genetic similarity, defined as average fraction of common fragments, by AFLP. Usual estimators are Dice (D) or Jaccard coefficients. D overestimates genetic similarity, since identical bands in profile pairs may correspond to different fragments (homoplasy). Another complicating factor is the occurrence of different fragments of equal length within a profile, appearing as a single band, which we call collision. The bias of D increases with larger numbers of bands, and lower genetic similarity. We propose two homoplasy- and collision-corrected estimators of genetic similarity. The first is a modification of D, replacing band counts by estimated fragment counts. The second is a maximum likelihood estimator, only applicable if band positions are available. Properties of the estimators are studied by simulation. Standard errors and confidence intervals for the first are obtained by bootstrapping, and for the second by likelihood theory. The estimators are nearly unbiased, and have for most practical cases smaller standard error than D. The likelihood-based estimator generally gives the highest precision. The relationship between fragment counts and precision is studied using simulation. The usual range of band counts (50–100) appears nearly optimal. The methodology is illustrated using data from a phylogenetic study on lettuce.

Introduction

AFLP is a DNA fingerprinting technique, that has been employed in many studies on plants (e.g. Tams et al. 2005), but also in studies on fungi (e.g. Mebrate et al. 2006), bacteria (e.g. Duim et al. 2001), and animals (e.g. Foulley et al. 2006). The resulting DNA fingerprints, also called profiles, are used in a wide spectrum of applications, like QTL studies (e.g. Zhong et al. 2006), diversity studies (e.g. van Berloo et al. 2008), and optimization of gene bank management (e.g. Jansen and van Hintum 2007). The question has been raised whether AFLP will remain useful in the near future, given the advances in genome sequencing, and new large-scale genotyping techniques like DArT (Wenzl et al. 2004). Meudt and Clarke (2007) suggest that fingerprinting techniques in general, and AFLP in particular, will remain valuable, especially if new analysis methods are developed, which overcome the problems arising in the analysis of AFLP data.

In this paper, we study the estimation of pairwise genetic similarity from dominant AFLP data. Estimation of similarity may be hampered by errors in, or erroneous interpretation of the binary band information from the AFLP profiles. As Bonin et al. (2007) mention, two types of errors prevail in AFLP genotyping: scoring errors and homoplasy. Many papers study the problem of scoring errors (e.g. parts of Meudt and Clarke 2007, and papers cited therein), but here we focus on homoplasy.

Estimation of genetic similarity is biased due to size homoplasy, see Fig. 1 (to be discussed later in greater detail). Size homoplasy occurs if, for two individuals, equally sized, but different DNA fragments comigrate in two AFLP lanes, resulting in identical bands. The two bands are usually considered homologous. Hence, part of the observed similarity can be attributed to chance. Size homoplasy is considered to be one of the major problems in the analysis of AFLP data (Meudt and Clarke 2007; Robinson and Harris 1999). Caballero et al. (2008) study the effect of size homoplasy on estimates of genetic diversity and detection of selective loci. Empirical estimates of the amount of homoplasy can be found, e.g. in O’Hanlon and Peakall (2000), who report that among congeneric thistles comigrating fragments showed on average 2.5% size homoplasy, but among different subtribes up to 100%. Because of this problem, AFLP is commonly advised to be used only to assess relationships of closely related taxa (Althoff et al. 2007).

Fig. 1
a Average Dice, and b average Jaccard similarities as a function of number of fragments for 100,000 simulated pairs of profiles with genetic similarities pgs = 0.0, 0.1, 0.3, 0.5, 0.7, 0.9, 0.95. Fragments are sampled from fld FS with ...

Another problem, related to the size homoplasy mentioned above, is the occurrence of two or more equally sized, but different fragments within a single lane. As two equally sized different fragments in two lanes generally comigrate, and are wrongly interpreted as homologous, they will also comigrate if amplified within a single lane, colliding in a single band, and wrongly interpreted as single fragment. We call the comigration of equally sized fragments within a single lane collision. In an empirical study on sugarbeet, Hansen et al. (1999) quantified the problem. They found 13.2% of the bands to contain collisions. In an in silico study of AFLP for a wide variety of species, Althoff et al. (2007) found fractions of bands containing collisions up to 49%, depending on the number of bands in a lane. Vekemans et al. (2002) reported in a Monte Carlo simulation study an average percentage of 30% of undetectable fragments. Collisions were studied from a probabilistic point of view in Gort et al. (2006) and Gort et al. (2008). Their theoretical results, which are at the basis of the present paper, are in line with the empirical results given above. Collisions also affect the estimation of genetic similarity.

Although it is recognized that both size homoplasy and collision may occur in AFLP, no attempts are usually made to correct for the problems: two equally sized bands are considered homologous, and a single band is interpreted as a single fragment. The reasons for this negligence are at least twofold: it is felt that the problems are minor (in the cases where AFLPs are suggested to be used), and hardly any methodology exists to correct for it. In Koopman and Gort (2004) a crude approach was proposed for the calculation of similarities from AFLP profiles.

In the present paper new estimators of genetic similarity from AFLP bands, corrected for homoplasy and collision, are proposed, one based on modification of the Dice and Jaccard coefficients, and one based on maximum likelihood. We take the following steps in the “Materials and methods” part to arrive at these estimators.

  • We first review the AFLP procedure as a sampling method of DNA fragments.
  • Next, the procedure and data are described from a modeling point of view, introducing notation, and a definition of pairwise genetic similarity for binary AFLP data is given.
  • We review some commonly used similarity coefficients.
  • We demonstrate, by simulation, that homoplasy and collision may seriously bias similarity estimates, resulting in Fig. 1.
  • A first step towards a solution is to estimate the number of fragments in a lane from the number of bands. We describe two ways to do this, depending on the availability of band position information.
  • Using estimated fragment counts, modified Dice (and Jaccard) coefficients in two versions are proposed, depending on availability of band position information.
  • If band position information is available, a second estimator of genetic similarity is proposed, based on maximum likelihood (m.l.).
  • Standard errors and confidence intervals are obtained, using the bootstrap for the modified coefficients, and standard likelihood theory for the m.l. estimator.
  • Further distributional characteristics of the estimators are studied by simulation. We describe precisely how we sample AFLP profiles.

Using the m.l. estimator and its precision, we next focus on the question how many bands in a lane should be used to estimate genetic similarity optimally. The theory is illustrated by a small case study on lettuce, using data from a phylogenetic study by Koopman et al. (2001). Results of the simulations and the case study are shown in “Results”. Conclusions are compiled and discussed in “Conclusions and discussion”. The paper ends with appendices on bootstrapping and an overview of all symbols used.

Materials and methods

AFLP reviewed

To understand the ideas we are proposing, a short review of the AFLP fingerprinting technique is useful. The AFLP technique, developed by Keygene N. V. (Vos et al. 1995), can be looked upon as a sampling technique of DNA fragments from, hopefully, random locations within a genome. To arrive at a sample of DNA fragments representing an individual genome four steps are taken:

  1. The total genomic DNA is cut into fragments by two restriction enzymes, often MseI (“frequent cutter”) and EcoRI (“rare cutter”). The result is a soup of fragments, flanked with restricted EcoRI–EcoRI, EcoRI–MseI, or MseI–MseI sites.
  2. Two adaptors, specific for the restriction enzymes, are ligated to the fragments, allowing primers to adhere in the third step.
  3. Two primers, complementary to the two adaptors, with one or more selective nucleotides select a number of fragments for PCR amplification. In this way a sample of fragments is drawn. Primers with more selective nucleotides will select fewer fragments. If the four nucleotides A–C–T–G occur equally often in the genome, one extra selective nucleotide on, e.g. the EcoRI primer will cause a fourfold reduction in sample size of EcoRI–MseI fragments, and a 16-fold reduction of the EcoRI–EcoRI fragments.
  4. The amplified fragments are separated by length in a lane of a gel or capillary electrophoresis system. Shorter fragments travel further. Usually only fragments with at least one EcoRI primer are labeled, and will become visible as bands. Only fragments with lengths within a certain scoring range (e.g. 51–500 nucleotides long) are visualized as bands.

On a single gel multiple individual genomes are fingerprinted, one per lane. The lengths of the bands are determined by comparison with the position of DNA fragments of known lengths (sizers) in size ladders. For a complete review of the AFLP technique, see e.g. Mueller and LaReesa Wolfenbarger (1999).

AFLP modeled: single profile

In this section, we again step through the AFLP procedure, but now aim to statistically model the procedure and data. For convenience, we compile all introduced symbols in Appendix 2 (Table (Table7).7). We describe the procedure for a single lane of a gel.

Table 7
Overview on symbols

In the first two steps of the procedure, the total genomic DNA is cut into fragments, and adaptors are ligated. Only part of these fragments are eligible for visualization: fragments containing at least one labeled site (e.g. Eco-RI site), and within the used scoring range (e.g. with 51–500 nucleotides) are candidates. We call this subset the population of fragments Π, containing, say, M fragments. Different restriction enzymes will result in different populations of fragments. The size and nucleotide composition of the genome also affect Π.

The length of a fragment is the number of nucleotides, adaptors included. We label the possible lengths of the fragments in Π with index i, ranging from 1 (referring to the smallest length in the scoring range) to N (referring to the largest length; e.g. with scoring range 51–500 N = 450). The probability distribution of the lengths is called the fragment length distribution (fld). With pi the probability that a fragment, randomly drawn from Π, has length i, we can write fld = (p1, p2,…, pN); note that equation M1 Shorter fragments are more frequent than longer fragments, i.e. the fld is monotonically decreasing and skewed to the right (Gort et al. 2006). The amount of skewness is mainly determined by the GC content of the genome, if the frequent cutter MseI is used. Lower GC content results in shorter fragments.

We assume the fld is known, or, at least, there is a reliable estimate of it. For all simulations we use fld FS, estimated from the Arabidopsis thaliana genome based on in silico AFLP, as in Gort et al. (2006). This fld is reasonable for genomes with GC content close to 36%. For the estimation of the fld for other genomes we refer to the same publication.

In step 3 the primers select a sample of fragments from Π, selecting only those fragments, which have specific nucleotides next to the restriction sites. This resembles systematic sampling, but with unknown sample size. We treat the lengths of the sampled fragments as a random sample from fld. Assuming a constant but unknown sampling probability π for the fragments of Π, the number of fragments in the sample, called k, has approximately a Poisson distribution with expected count m = πM.

In step 4 the k fragments are separated by length, and visualized as bands. We assume that the position of a band within a lane is determined principally by the fragment length. Hence, a band will occur approximately at one of N discrete positions within a lane, which we call band lengths. A consequence is that two different fragments of the same length will occur as a single band.

The end product is a profile, containing bands at discrete positions, which can be represented by a binary vector y = (y1, y2,…, yN). The binary variable yi (i = 1,…, N) indicates whether a band with length i is present. The number of bands in a lane is equation M2 Notice that the number of bands cannot be larger than the number of fragments (n ≤ k).

AFLP modeled: pairs of profiles and their similarity

Two related individuals share parts of their DNA. As a consequence, they share part of their two populations of fragments Π1 and Π2, containing M1 and M2 fragments, formed at step 2. This common part is called Πa, and contains Ma fragments. The complement of Πa within Π1 is called Πb, consisting of Mb fragments present in individual 1, but absent in 2. The complement of Πa within Π2 is called Πc, and consists of Mc fragments, present in 2, but absent in 1. Πb and Πc are called the populations of unique fragments. Notice that M1 = Ma + Mb, and M2 = Ma + Mc. All population sizes Ma, Mb, and Mc are unknown. The fractions of common fragments are F1 = Ma/M1 and F2 = Ma/M2, which need not be the same, e.g. if the genomes have different sizes.

We define the pairwise genetic similarity pgs of a pair of genotypes as the weighted average of fractions F1 and F2, with weights proportional to the population sizes:

equation M3
1

Notice that pgs can be written as 2 Ma/(2 Ma + Mb + Mc).

We assume that Πa, Πb, and Πc have the same fragment length distribution fld.

In step 3 samples from fld are taken, resulting in sample sizes of fragments ka, kb, and kc, approximately Poisson distributed with expected fragment counts ma, mb, and mc, proportional to Ma, Mb, and Mc. The expected numbers of fragments of the two profiles are m1 = ma + mb and m2 = ma + mc.

The end product after step 4 is a pair of profiles, represented by two binary band vectors y1 = (y11,…, yN1), and y2 = (y12,…, yN2), with band counts equation M4 (j = 1, 2). We use the following notation for band counts:

  • a = number of shared bands in the two profiles = equation M5;
  • b = number of bands in the first profile, but absent in the second = equation M6;
  • c = number of bands in the second profile, but absent in the first =equation M7;
  • d = number of empty positions in both profiles = equation M8.

Hence, a, b, c and d are the number of 1–1, 1–0, 0–1, and 0–0 matches, respectively. If more than two profiles are compared, d is often defined as the number of 0–0 matches in two lanes, limited to those band lengths with at least one band in one of the other lanes.

Commonly used similarity coefficients

We now review some commonly used similarity coefficients for binary AFLP data. From the similarity coefficients, reviewed by Reif et al. (2005), only the Dice, Jaccard’s, and simple matching coefficient are relevant, because we treat AFLP as a dominant marker system.

The Dice coefficient (Dice 1945) D is an estimator of pgs: equation M9 with weights equation M10, equation M11, and equation M12, equation M13. In genetic contexts the Dice similarity is often referred to as the Nei–Li similarity (Nei and Li 1979).

The Jaccard coefficient (Jaccard 1908) equation M14 is the fraction of common bands compared to the total number of different bands for the two profiles. It is an estimator of Ma/(Ma + Mb + Mc), and not of the genetic similarity, as we define it. A non-linear relationship exists between J and D: equation M15. For example, taking equal band counts in the two profiles: if half of the bands in each profile is shared, then D = 1/2, and J = 1/3. Examples of applications of Dice and Jaccard’s coefficients as measures of genetic similarity are Drossou et al. (2004), and Tams et al. (2005).

The simple matching coefficient (Sneath and Sokal 1973) equation M16 measures similarity including the 0–0 matches in the profiles as well, counting the 1–1 and 0–0 matches alike.

To illustrate the differences between the coefficients, take two genotypes with profiles containing 100 bands each, with N = 450, a = 50, b = 50, c = 50, hence d = 300. Since half of the bands of each profile is shared, D = 0.5, and J = 0.33, whereas S = 0.78. Suppose that for the same genotypes a second set of profiles is made, using primers with more selective nucleotides, and hence smaller samples of amplified fragments. Assuming a proportional decrease of band counts of 50% (so a = 25, b = 25, c = 25, and d = 375), we still have D = 0.5, and J = 0.33, but S = 0.89. Hence, S changes if the band counts decrease proportionally, whereas D and J remain constant.

Usually more than two genotypes are compared in a study. Often, for S only the 0–0 matches are counted for the occupied band positions in the whole set of genotypes. With a proportional decrease of the band counts a, b and c, the null count d will also decrease, but likely at a different rate. Hence, S will likely change, whereas D and J remain constant. S can also change if the set of other genotypes under study is changed. Wong et al. (2001) supply reasons in the realm of codominance of AFLP to avoid similarity measures exploiting 0–0 matches. Therefore, S has a number of undesirable properties. Only D is an estimator of pairwise genetic similarity, as we have defined it.

The problem: homoplasy and collision

To appreciate the possible consequences of homoplasy and collisions in relationship studies based on AFLP data, we performed a simulation study. We sampled 100,000 pairs of profiles for a range of genetic similarities pgs (=0, 0.1, 0.3, 0.5, 0.7, 0.9, 0.95) and fragment counts m1 = m2 (=1,…, 200). The maximum fragment count m = 200 corresponds to ≈140 bands, which is about the maximum number of bands per lane to be found in practice. Each pair was sampled in three steps. First, a random draw ka from the binomial (m1, pgs) distribution determined the sample size of fragments from the common part Πa, the remaining kb = m1 − ka and kc = m2 − ka fragments to be sampled from the unique parts Πb and Πc. Next, ka, kb, and kc lengths were sampled from the fld, and results were combined into two vectors of length N = 450, containing the counts of lengths 1,…, 450 for the two profiles. In the last step, a pair of binary vectors was created, containing absence/presence information of at least one fragment of length 1,…, 450, and representing a pair of AFLP profiles. Dice and Jaccard coefficients D and J were calculated for each pair, and averaged over all pairs to produce Fig. 1. The graph shows the average D and J as a function of the fragment count. The average band count is shown at the top axis on a non-linear scale. As an example, profiles with 100 fragments tend to produce approximately 83 bands, hence 17 collisions.

D overestimates the true genetic similarity seriously, increasingly so for larger fragment or band counts, and for smaller genetic similarities. For example, at band count 60 the average D has approximate biases 0.015, 0.085, and 0.23 for pgs = 0.9, 0.5, and 0.0, respectively. At band count 100 the biases are 0.025, 0.14 and 0.34, respectively.

J is for band counts in the range 60,…, 100 sometimes lower than the true pgs (if pgs > 0.3), sometimes close to pgs (if pgs ≈ 0.3) and sometimes higher (if pgs < 0.3).

Estimation of number of fragments

The basic idea in this paper is that, in order to estimate genetic similarity, we need to know how many fragments from the two profiles are identical, whereas the profiles indicate how many bands are identical. The first step to solve this problem is to estimate the expected number of fragments m that gave rise to the n observed bands in a single profile. The difference between number of fragments and number of bands is called the collision count.

To estimate m, we discriminate between situations without and with band length information. Notice that band lengths are not always available, although in principle the information can be read from an AFLP gel, if size ladders are used. The lack of band length information is often based on limitations in the realm of intellectual property, as commercial players like Keygene N.V. propagate.

In the case of unknown band lengths, the collision count for a given fld is estimated from the band count, using Bayes’ rule and generalized occupancy distributions, see Gort et al. (2006). The resulting estimator of the expected number of fragments m is called equation M17.

With known band lengths, the number of collisions can be estimated using Bayes’ rule and approximated multinomial tail probabilities, or applying the EM-algorithm, as in Gort et al. (2008). In the present paper, we report a simpler approach to arrive at an estimator of m. We propose a generalized linear model (g.l.m.) (McCullagh and Nelder 1991) for the binary band scores yi. The scores yi are assumed to be independent, and Bernoulli (Pi) distributed, with expected score E(yi) = Pi the probability that a band occurs with length i, if a sample of m fragments has been drawn from fld=equation M18. The band probability Pi and fragment probability pi are related as: equation M19, because the event “no band of length i” is equivalent with “none of the m fragments has length i”. This equation can be transformed into

equation M20

revealing the systematic part of the g.l.m. Hence, we fit a regression model for the band scores yi, using log(m) as intercept, offset equation M21, and complementary log–log link. The estimator equation M22 of m is obtained by exponentiation of the estimator of the intercept log(m).

Modified Dice and Jaccard coefficients using binary AFLP data

Suppose we have two profiles with observed band counts n1 = a + b, and n2 = a + c. The expected numbers of fragments m1 and m2 are estimated by equation M23 and equation M24 by either of the two estimators from the previous section. The pairwise genetic similarity to be estimated is equation M25, as in (1).

For weights w1 and w2, we have straightforward estimators equation M26, and equation M27, since expected fragments counts are assumed to be proportional to population sizes. However, for the fractions common fragments equation M28 and equation M29, an estimator equation M30 of the number of common fragments ma is needed. We estimate ma as equation M31, by analogy with the number of shared bands a, which can be calculated as a = n1 + n2 − n1+2. In this formula n1+2 = a + b + c is the total number of different bands, as if combining the two profiles into a single profile, and counting the bands. In the formula for equation M32, equation M33 is the estimated fragment count for the combination of the two profiles. The rationale of estimator equation M34 is the following: equation M35 estimates the number of fragments from the n1 bands of profile 1, and equation M36 from the n2 bands of profile 2. The sum equation M37 estimates the total number of fragments in the two lanes. Some of the fragments are counted twice, as they occur in both profiles. If we overlay profiles 1 and 2, we see what would have happened if we mixed the populations of fragments for the two genomes, and made a profile for the mixture. Identical fragments in the two populations, selected for amplification, will appear as a single band now, and equation M38 estimates the total number of fragments in the profile for the mixture, that is the number of different fragments in the mixture. Then the difference (equation M39) – equation M40 estimates equation M41, i.e. the number of fragments the two profiles have in common.

This results in equation M42 and equation M43. Estimators of unique fragment counts are equation M44, and equation M45.

As estimator of genetic similarity pgs we now propose the modified Dice coefficient

equation M46

replacing the band counts in the original Dice coefficient by estimated fragment counts. The Jaccard coefficient may be modified in the same way:

equation M47

The maximum of both Dmod and Jmod is 1, occurring if the two profiles are identical. At the other end of the scale, there is no intrinsic limitation both for Dmod and Jmod to take on negative values, whereas pgs ≥ 0. A solution to the problem is truncation of the estimator at 0.

The modified coefficients come in two versions, for situations without and with band length information. If band lengths are unknown, estimator equation M48 is used, resulting in modified Dice and Jaccard coefficients

equation M49

If band lengths are known, we use estimator equation M50, and the modified coefficients become

equation M51

Maximum likelihood estimator of genetic similarity from binary AFLP data

In the case of known band lengths, a second estimator equation M52 of the genetic similarity pgs is proposed, based on maximum likelihood (m.l.) (Silvey 1975). For this estimator we need a statistical model for the data, consisting of the N pairs of binary scores equation M53, equation M54 We treat these pairs as independent. The two profiles have expected fragment counts equation M55 as before.

The four possible outcomes of a pair equation M56 are:

  1. (0,0): no fragment of length i at all;
  2. (0,1): no fragment from the unique part Πb of genotype 1 and the common part Πa, and at least one fragment from the unique part Πc of genotype 2;
  3. (1,0): at least one fragment from Πb, and no fragment from Πc and Πa;
  4. (1,1): either at least one fragment from Πa, or at least one fragment from both Πb and Πc, but not from Πa.

The likelihood of these four outcomes for the ith pair is:

  1. (0,0): equation M57
  2. (0,1): equation M58
  3. (1,0): equation M59
  4. (1,1): equation M60

Next, the log-likelihood of the data equation M61 is maximized with respect to the parameters ma, mb, and mc, resulting in m.l. estimators equation M62, equation M63, and equation M64. As in the previous section, we can define a modified Dice coefficient, now based on m.l. estimators, as

equation M65

with weights equation M66, equation M67, and equation M68, equation M69.

The m.l. procedure returns approximate standard errors of equation M70, equation M71, and equation M72, but not of equation M73 as an estimator of pgs. To get the precision of an estimator of pgs, we reparameterize the likelihood. From equation M74, it follows equation M75, since we assume expected fragment counts proportional to population counts. Now, we replace ma in the likelihood above by equation M76. Now the log-likelihood is maximized with respect to pgs, mb, and mc, resulting in a direct m.l. estimator of pgs, which we call equation M77.

A third parameterization replaces ma by equation M78, with lgs = logit(pgs), yielding an estimator on the logit-scale, to be back-transformed to equation M79. This estimator may have better distributional properties for pgs close to 0 or 1.

Precision of the estimators

The precisions of estimators equation M80 and equation M81 are determined by bootstrapping (Efron and Tibshirani 1993), whereas for equation M82 the precision follows from standard likelihood theory.

For estimator equation M83 the following bootstrap method is used. The data for a pair of profiles consists of a pairs 1–1, b pairs 1–0, c pairs 0–1, and d pairs 0–0, collected in the vector (a, b, c, d), without knowledge of band lengths. For one bootstrap resample we take a sample of size N from the pairs 1–1, 1–0, 0–1, and 0–0, with probabilities given by a/N, b/N, c/N, and d/N, respectively. For this bootstrap sample the modified Dice coefficient is calculated as described, and stored.

For estimator equation M84 a different bootstrap method is used. Now the band lengths are known. A bootstrap resample consists of a sample with replacement of N pairs equation M85 and connected fld probabilities pi from the N pairs equation M86, equation M87 and a rescaling of the set of pi’s to have sum 1. Notice that the same pair equation M88, i.e. with the same band length, may occur more than once in the bootstrap resample. Therefore, a single bootstrap resample does not necessarily correspond to a pair of profiles, which could occur in practice. The method nevertheless works well, as shown later.

For equation M89 and equation M90 we took 1,000 bootstrap samples, resulting in estimates of bias (defined as bootstrap mean − estimate), standard error, and bootstrap confidence intervals. We used accelerated bias-corrected percentile bootstrap confidence intervals, also known as BCa confidence intervals (DiCiccio and Efron 1996). For a description of the calculation of these confidence intervals, as well as a comparison between different types of bootstrap confidence intervals, we refer to the appendix.

For estimator equation M91 approximate standard errors follow from standard likelihood theory, leading to Wald confidence intervals for pgs as equation M92, with z1-α/2 the 1 − α/2 quantile from the standard normal distribution. For equation M93 we back-transform the Wald-confidence interval equation M94 using equation M95. Besides Wald-type confidence intervals we calculated profile likelihood confidence intervals for pgs (see e.g. Venzon and Moolgavkar 1988). For profile likelihood confidence intervals the parameters mb and mc are treated as nuisance parameters, resulting in a profile likelihood for pgs by maximizing over mb and mc.

Sampling of AFLPs and simulation

To study the behavior of the proposed estimators, we performed a simulation study. For a wide range of parameter settings (pgs, m1, and m2) pairs of profiles were simulated by

  1. calculating the expected counts of common fragments ma = ½(m1 + m2)pgs, and unique fragments mb = m1 − ma, and mc = m2 − ma;
  2. drawing random counts from Poisson distributions with means ma, mb, and mc to arrive at fragment counts ka, kb, and kc for the pair of profiles to be generated;
  3. sampling separately ka, kb, and kc fragment lengths from the fld;
  4. combining the ka + kb sampled fragments into the first profile, and ka + kc fragments into the second, condensing the information into binary vectors y1 and y2 of length N.

For all combinations of pgs = (0, 0.1, 0.3, 0.5, 0.7, 0.9, 0.95) and m1 = m2 = (40, 70, 120), we sampled 10,000 pairs of profiles. We also included a selection of unequal m’s for some values of pgs, to show that the methodology works in that case as well. For each pair of profiles the estimates equation M96, equation M97 (with 1,000 bootstrap samples), and the three versions of equation M98 were calculated.

Application of methodology: effect of number of fragments on precision

In AFLP profiling the number of fragments in a lane, and hence the number of bands, can be steered by the researcher by changing the number and/or type of selective nucleotides of the primers. Typical band counts per lane are between 50 and 100, corresponding to fragment counts from 60 to 125. The question arises whether these typical counts are optimal, i.e. whether the estimators of genetic similarity have highest possible precision.

In a simulation study we investigated for a number of examples (as before, pgs = 0.0, 0.1, 0.3, 0.5, 0.7, 0.9, and 0.95 using N = 450 and fldFS), how the standard error and width of the 95% profile likelihood confidence interval of pgs based on equation M99 depends on the fragment count. Expected fragment counts were varied from 15 to 500 (in steps of 5, equal expected counts for pairs of profiles), using 10,000 replicates at each step. We pushed the number of fragments to unrealistically high values now, to show the properties of equation M100 in that case, at the same time realizing that in practice it is impossible to score profiles with very large numbers of bands per lane. In the simulations numbers of fragments up to 500 were allowed, resulting in profiles with more than 225 bands on average. In that case more than half of the band positions are occupied, since N = 450.

Case study: phylogenetic relations between Lactuca genera

The lettuce study by Koopman et al. (2001) aims at inferring species relationships in Lactuca and related genera from AFLP fingerprints. We selected one of the two primer combinations (E35/M49), and only 5 of the 20 species: L. tenerrima, M. muralis, L. serriola, L. sativa, and L. tatarica. We took 6–9 accessions for each of the five selected species. We selected the five species to have a wide range of band counts: mean counts (±SD) are 29.6 (±1.9), 32.4 (±2.5), 49.6 (±3.0), 52.6 (±2.8), and 84.1 (±5.1) for L. tenerrima, M. muralis, L. serriola, L. sativa, and L. tatarica, respectively.

For all pairs of accessions we calculated D, J, and Dmle. We used FS from A. thaliana as fld. This seems reasonable, since the GC content of lettuce is close to that of A. thaliana: 36.6, 37, 38.2, 38.3, and 36.4% for the five species (Koopman et al. 2002) versus 36% for A. thaliana. The relationships between the species are visualized with UPGMA dendrograms, using dissimilarities 1 − D, 1 − J, and 1 − Dmle.

Results

General results from the simulation study

Table 1 shows some general results from the simulation study. For all simulation settings of pgs, m1, and m2, the average band counts n1, n2, and average Dice similarity D are given. From the comparison of expected fragment counts with average band counts, we note that profiles with m = 40 have on average three collisions, with m = 70 on average 8.7 collisions, and with m = 120 on average 23.6 collisions. The ordinary Dice coefficient seriously overestimates the true similarity, with largest biases for small similarities and large fragment counts. The maximum observed bias is 0.334 for pgs = 0 and m = 120. The smallest bias is 0.0034 for pgs = 0.95 and m = 40.

Table 1
Average band counts n1 and n2, and Dice similarities D for 10,000 simulated pairs of AFLP profiles for a range of values of genetic similarity pgs and expected numbers of fragments m1 and m2

Results from the simulation study for modified Dice coefficients

Table 2 shows the results from the simulation study for the modified Dice coefficient equation M101, using profiles without band length information. In Table 3 results for equation M102 are given. We notice the following.

  1. Almost all of the bias of the original Dice coefficient is removed. equation M111 and equation M112 slightly underestimate pgs now (mean observed biases −0.0018 and −0.0015, averaged over all settings of pgs and m, for equation M113 and equation M114, respectively), with largest observed bias equal to −0.0030 occurring for equation M115 in case pgs = 0 and m = 120. The remaining small negative bias can be removed even further by using a bootstrap bias correction. Mean observed biases are then −0.00058 and −0.00025.
  2. The 95% (BCa bootstrap) confidence intervals for the genetic similarity pgs show reasonably good coverage properties. In 21 and 18 out of the 25 experimental settings the observed non-coverage is between 4.5 and 5.5%, hence deviations less than 0.5% from the nominal value of 5%. For both estimators the largest deviation from 5% is found for pgs = 0.90 and m = 40, with observed non-coverages of 3.8 and 3.8%, respectively. In these cases the confidence intervals are slightly too wide. For pgs = 0.95 and m = 40 the overall non-coverage behaves better (5.2 and 5.3%), but we find that in 1.6 and 1.7% of the cases the confidence intervals are too low, and in 3.6 and 3.5% too high, compared to the nominal 2.5 and 2.5%. In this case the intervals are too wide if the estimate is smaller than pgs = 0.95, and too narrow for estimates larger than 0.95.
  3. The bootstrap standard errors of equation M116 and equation M117 are smaller for larger number of expected fragments, with the exception of pgs = 0 and pgs = 0.1. Hence, in the examples for pgs > 0.1 larger fragment counts result in more precise estimates. The same can be said for the lengths of the 95% confidence intervals. If pgs = 0.1 the smallest standard error is observed for m = 70
  4. The estimates equation M118 and equation M119 may become negative for small values of pgs. In the table this can be seen for pgs = 0, resulting in a negative average of Dmod, but it also occurs for pgs = 0.1. For pgs = 0.3 the lower bound of the 95% confidence interval may become negative. In practice a negative value of Dmod would be truncated at 0. Therefore, we added the bottom parts II of Tables 2 and and3,3, showing results for the truncated versions of equation M120 and equation M121 for pgs = 0.0, 0.1, and 0.3. Since the truncation causes more distributional asymmetry we give medians instead of averages of equation M122 and equation M123. For equation M124 the bias-correction decreases the bias, but this is not always the case for equation M125. For pgs = 0 we give the non-coverage of the (97.5%) confidence interval only at the right of pg = 0. For pgs = 0 we observe the largest standard errors for the cases with largest m, suggesting that the optimal number of fragments is smaller than m = 120.
  5. In all cases equation M126 has narrower 95% confidence intervals than equation M127, although differences are small (average difference in length is only 0.0019). In all cases the bootstrap SE(equation M128) ≤ SE(equation M129), but again differences are small. The coverage of the 95% confidence interval of equation M130 is slightly better than that of equation M131: average absolute deviation from the nominal 5% is 0.33% for equation M132 compared to 0.36% for equation M133. Intuitively better behavior of equation M134 was expected, since equation M135 exploits band length information, but we conclude, surprisingly, that equation M136 has slightly better characteristics than equation M137
Table 2
Results from a simulation study on equation M103 for a range of values of genetic similarity pgs and expected numbers of fragments m1 and m2, 10,000 replicated pairs of AFLP profiles, 1,000 bootstrap resamples, fldFS from A. thaliana with N = 450
Table 3
Results from a simulation study on equation M107 for a range of values of genetic similarity pgs and expected numbers of fragments m1 and m2, 10,000 replicated pairs of AFLP profiles, 1,000 bootstrap resamples, fldFS from A. thaliana with N = 450

Results from the simulation study for maximum likelihood estimators Dmle

Table 4 shows the results from the simulation study for equation M138. We notice the following.

Table 4
Results from a simulation study on equation M139 for a range of values of genetic similarity pgs and expected numbers of fragments m1 and m2, 10,000 replicated pairs of AFLP profiles, 1,000 bootstrap resamples, fldFS from A. thaliana with N = 450
  1. Estimators equation M147, equation M148, and equation M149 almost always return the same estimate. Only for pgs ≥ 0.9 we see minor differences, resulting in means differing in the fourth decimal. Hence, only results for equation M150 are shown.
  2. The large positive bias of the original Dice coefficient is removed. For pgs > 0.1, a negligible negative bias of equation M151 remains: the mean bias is −0.0015. For pgs ≤ 0.1 a small positive bias is observed, because of the necessarily non-negative value of the estimators. For pgs = 0 the medians (not shown) are 0, and for pgs = 0.1 they are 0.0965 (m = 40), 0.0982 (m = 70), and 0.0995 (m = 120).
  3. The 95% Wald confidence intervals for pgs are conservative for small values of pgs (non-coverage rates smaller than nominal value), but are becoming more and more liberal for larger values. Obviously, the approximate standard error of equation M152 is too large for small values of equation M153, and too small for large values. The deviations from 5% seem acceptable for 0.3 ≤ pgs ≤ 0.7 and m > 40. The number of intervals with a lower bound larger than the true pgs outnumber those with an upper bound smaller than pgs. This is also an indication of standard errors which are too high for low values of the estimate, and too small for large values.
  4. The 95% profile likelihood confidence intervals for pgs have for a large number of settings non-coverage rates close to 5%. In 16 out of the 25 settings the deviation of the non-coverage rate from the nominal value is less than 0.5%. Larger deviations are found for larger values of pgs and smaller fragment counts. The largest deviation is observed for pgs = 0.95 and m = 40, with a non-coverage rate equal to 19%, making the profile likelihood interval useless in this situation. The number of intervals with an upper bound smaller than pgs becomes exceedingly large in these cases. The profile likelihood intervals work well for pgs < 0.7, irrespective of the studied fragment counts, and for larger values of pgs, but only if the fragment count is large enough.
  5. The 95% back-transformed (from logit-scale) Wald confidence intervals generally have a non-coverage rate close to the nominal 5%. However, for small values of pgs they are highly asymmetrically distributed (with respect to pgs). Intervals with lower bounds exceeding pgs dominate in these cases. If pgs = 0, estimates of pgs on the logit scale tend to −∞, and the approximate standard errors are badly determined, resulting in useless confidence intervals. For high values of pgs, intervals with upper bounds lower than pgs get the upper hand. The back-transformed Wald confidence intervals are usable for pgs ≥ 0.5, and tend to be conservative then.
  6. The standard error of equation M154 decreases with larger expected fragment counts, as expected. For all three types of confidence intervals larger numbers of fragments result in narrower confidence intervals.
  7. None of the three types of confidence intervals are usable for all values of pgs. The profile likelihood intervals have the broadest range of application of pgs:pgs < 0.7 irrespective of m, and pgs ≥ 0.7 for larger values of m. The back-transformed Wald intervals perform best for large values of pgs. The Wald confidence intervals are widest (at pgs = 0.5 and 0.7), making them the least attractive in this range.
  8. For all cases with pgs ≥ 0.3, equation M155 has smaller standard errors than equation M156 and equation M157. Furthermore, in all cases the profile likelihood confidence intervals based on equation M158 are narrower than the bootstrap confidence intervals based on equation M159 and equation M160. These results suggest that equation M161 is to be preferred over the modified coefficients equation M162 and equation M163.

Comparing standard errors

The simulation study has shown that the proposed estimators are approximately unbiased. Although attractive in itself, unbiasedness does not guarantee a higher precision since equation M164. Using the data from the simulation study, we estimated equation M165, and equation M166 by bootstrapping, and compared equation M167 with equation M168.

For most cases we find equation M169 > equation M170, with the most extreme outcome for pgs = 0.0 and m = 120, where equation M171 is equation M172. For large values of pgs (pgs = 0.95, all m; pgs = 0.9, m = 40,70; pgs = 0.7, m = 40), we find that equation M173 < equation M174, but equation M175 is never smaller than equation M176. Hence, depending on the combination of pgs and m, very large gains in standard error can be obtained, or, for large pgs (in combination with small fragment counts) minor losses. In the last cases, the gain in bias is outweighed by the loss in variance, and the new estimator equation M177 is marginally less precise compared to D.

Results for the effect of expected number of fragments on precision

Figure 2 shows the results of the simulation study on the relationship between the expected number of fragments m and precision of equation M178. In the left-hand side figure the expected number of fragments is plotted against the average standard error of equation M179. At the top axis the average band count is shown. We observe the following:

  1. Starting at small numbers of fragments, the standard error of equation M180 decreases as the number of fragments increases. The rate of change of the standard error is high at low fragment counts, but decreases. As the number of fragments increases, the standard error reaches a minimum, and afterwards increases again.
  2. The optimal number of fragments depends on pgs. Smaller values of pgs allow smaller numbers of fragments. For pgs = 0 or 0.1 the optimal number of fragments is close to m = 140 (or n = 110 bands). For pgs = 0.3 this count is approximately m = 250 (n = 165), for pgs = 0.5 m = 350 (n = 205), and for pgs = 0.7 m = 500 (n = 245). For pgs = 0.9 or 0.95 the optimal fragment count is larger than 500 fragments.
  3. In general a large range of near-optimal fragment counts exists.
  4. The usual range of band counts (between 50 and 100) is not optimal, especially if the focus is on highly related species with high pgs. However, the gain in accuracy will generally be small if larger band counts are used. The small gain in accuracy must be balanced against the possible scoring problems that may occur with large band counts.
Fig. 2
a Average SE of equation M183, and b average equation M184 and D, as functions of numbers of fragments for different values of pgs. In plot a interpolated lines are drawn for fragment counts ranging from 15 to 500 in steps of 5 for pgs = 0, 0.1, 0.3, 0.5, 0.7, ...

In the right-hand side figure the expected number of fragments is plotted against the average equation M181, and average Dice similarity. Furthermore, the average lower and upper bounds of the 95% profile likelihood confidence intervals are shown. For clarity, only results for pgs = 0.1, 0.5, and 0.9 are given. We observe the following:

  1. equation M182 is an (almost) unbiased estimator of pgs, even for extremely large fragment counts. For very small fragment counts (m ≤ 25) there appears to be small negative bias.
  2. Starting at small m, the width of the confidence interval quickly decreases. For large enough m (depending on pgs) the width remains approximately constant.
  3. The usual range of band counts, although not optimal, seems reasonable. Only little gain in the width of the confidence intervals can be expected from higher fragment counts, as in 4).
  4. The confidence intervals are rather wide. The only way to reach narrower intervals is to use multiple gels with different primer combinations, and combine the information from the different profiles.

Results for case study on lettuce and related genera

Figure 3 shows the UPGMA dendrograms for the five species, split out for the three dissimilarity measures. The dendrograms for 1 − D and 1 − J are largely the same. With all three dissimilarities the species are separated well. Notice that the 1 − Dmle dissimilarities are closer to 0, as expected. Notice further that the 1 − Dmle dissimilarities are not a simple shift towards 1. In the hierarchical clustering scheme for D and J, L. tenerrima joins after clustering of L. serriola, L. sativa, and L. tatarica, but for DmleL. tenerrima joins after clustering of L. serriola and L. sativa only. Apparently, L. tenerrima and L. tatarica have switched places. This behavior can be understood from the band count. The AFLP profiles for L. tenerrima contain a small number of bands, whereas L. tatarica profiles have large counts. Hence, bias corrections for comparisons with L. tenerrima are smaller than those with L. tatarica.

Fig. 3
UPGMA dendrograms for three dissimilarities: a 1−Dice D, b 1−Jaccard J, and c 1−Dmle for five species of Lactuca and related genera, with 6–9 accessions per species. Labels are: ten = L. tenerrima, mur =  ...

Conclusions and discussion

In this study, we propose new estimators of pairwise genetic similarity pgs from binary AFLP data, correcting for homoplasy. We define pairwise genetic similarity for AFLP data as the weighted average of fractions of common fragments. Using this definition, the Dice coefficient is a natural candidate for replacement, but a homoplasy corrected version of the Jaccard coefficient is suggested as well. For most practical cases the new estimators are better than the ordinary Dice coefficient, because the bias is removed, at the cost of a small increase in variance. Only for large genetic similarities in combination with low band counts (roughly: pgs = 0.95 and n < 100, pgs = 0.90 and n < 65, pgs = 0.70 and n < 38), Dice performs better.

For profiles without band length information, we propose the modified Dice coefficient equation M185. Using the bootstrap, standard errors and confidence intervals are obtained. The bootstrap allows a further reduction of the already small negative bias of equation M186.

For AFLP profiles with band length information, we have three candidate estimators: equation M187, equation M188, and equation M189. Best results are obtained using the maximum likelihood estimator equation M190, although differences are small. Second best is, surprisingly, equation M191, ignoring the band length information. The standard error of equation M192 follows from likelihood theory, hence no bootstrapping is needed. Profile likelihood confidence intervals for pgs are narrowest. However, care has to be taken in the choice of type of confidence interval. Profile likelihood intervals are only acceptable, if pgs < 0.7 irrespective of the number of fragments, and for pgs ≥ 0.7 if the fragment counts are large enough. For small fragment counts and large pgs, more acceptable results are obtained for the back transformed Wald intervals, using an estimator on the logit-scale.

The modified Dice coefficients equation M193 or equation M194 are good alternatives as well. Over the whole range of pgs the confidence intervals based on equation M195 and equation M196 showed more stable coverage properties than those for equation M197.

The homoplasy corrected estimate of genetic similarity is always smaller than the ordinary Dice coefficient, because part of the observed band similarity is attributed to chance. The magnitude of this correction depends on the true genetic similarity, but also on the fragment counts. Both smaller similarities and larger numbers of fragments lead to larger corrections.

The standard error of the similarity estimator equation M198 and the width of the confidence interval cannot be made arbitrarily small by increasing the number of fragments in the profiles. The optimal number of fragments exists, but its value depends on the true genetic similarity, and there is a large range of near-optimal fragment counts. The usual range of band counts (between 50 and 100) is suboptimal, but in general the gain in precision is small if higher numbers of fragments are used, and should be balanced against increasing scoring problems.

To get more precise estimates of genetic similarity, multiple gels with different primer combinations or restriction enzymes should be used, and the information from the different profiles should be combined. equation M199 can easily be modified to estimate a single genetic similarity from multiple pairs of profiles, even allowing for possibly different fld’s for the different profiles. Modifications of this type (beyond ordinary averaging) are less straight forward for the modified coefficients equation M200 and equation M201. This flexibility is a further argument in favor of equation M202.

To account for homoplasy and collisions properly, all bands in the profiles must be scored, not just the non-monomorphic bands. The effect of scoring non-monomorphic bands only is that Dice and Jaccard coefficients are lowered in a way that depends on the set of individuals under study. Inclusion or exclusion of a less related individual in the study, could result in exclusion or inclusion of bands, which are polymorphic with the individual, but monomorphic without. Hence, the similarity coefficient would be different with or without this individual.

Conclusions drawn here are mainly based on a single simulation study. Furthermore, we have to rely on a number of assumptions. For instance, we assume to know the fld, which in reality hardly ever is the case. Only if full DNA sequence information is available and by using in-silico AFLP procedures, do we have an estimate of the fld very close to the true fld. In other cases, a less reliable estimate of the fld may come from the GC content or directly from the binary AFLP data, as described in Gort et al. (2006).

Another topic related to the fld, is the fact that two distantly related individuals, e.g. with highly different GC contents, may have different flds. In this paper we have assumed that there is a common fld. Further study on the effect of misspecification of the flds on the statistical properties of the proposed estimators is needed.

In the present paper, we studied the effect of homoplasy and collision on the estimation of genetic similarity from binary AFLP data. Examples of studies that may directly benefit from the proposed homoplasy corrected estimates of genetic similarity are studies on genetic diversity, e.g. in plant genetic resources or breeding programs, but also phylogenetic and taxonomic studies, and studies of essential derivation, in which plant breeders try to establish thresholds for genetic similarity between initial and new, allegedly derived varieties (Van Eeuwijk and Law 2004).

In other studies where AFLP profiles are analyzed, the problem of homoplasy may have an impact as well. For example, in linkage studies for tracing quantitative trait loci (QTLs) or for mapping purposes, a band is interpreted as a single DNA fragment, residing at one unique locus of the genome. Here the best strategy may be to avoid homoplasy as much as possible, by limiting the number of fragments per lane, or avoiding bands corresponding to short fragments.

In population genetic applications of AFLP, homoplasy and collision may also affect estimation of parameters. For example, if the allele frequency of the DNA fragment corresponding to a band is the parameter of interest, like in Krauss (2000), who tested three procedures for estimation of null allele frequencies, homoplasy may cause some bands to be non-homologous, thereby changing the relative frequency of absent bands. Derived quantities like heterozygosity, coefficient of co-ancestry, or genetic distances, may need corrections for homoplasy and/or collision as well. These corrections require careful consideration, and are beyond the scope of the present paper. An example of a recent study of homoplasy in population genetics is Caballero et al. (2008), who focus on population genetic diversity and detection of selective loci.

In a study by Holland et al. (2008) about automated scoring of AFLPs, the suggestion is made to decrease the bin width for scoring fragments on a capillary system. This is another route towards a solution of the homoplasy problem, because the resulting profiles will likely have less homoplasy, albeit at the cost of an increased error rate for homologous fragments. In future work this approach may be joined with ours to arrive at improved evaluation of homoplasy.

The problem of homoplasy described here is not limited to the AFLP marker system. In a study on homology among RAPD fragments for three very closely related species of sunflowers, Rieseberg (1996) reports that of 220 pairwise comparisons of comigrating fragments only 79% identified loci useful for comparative genetic studies. For RAPD comparable corrections for homoplasy can be envisioned, as we propose here for AFLP.

Software in R (R Development Core Team 2005) for calculation of the proposed estimators is available from the authors.

Acknowledgments

We thank Wim Koopman for providing the data from the lettuce study, Herman Adèr for discussing bootstrap procedures, and Alfred Stein for proof reading the manuscript.

Open Access This article is distributed under the terms of the Creative Commons Attribution Noncommercial License which permits any noncommercial use, distribution, and reproduction in any medium, provided the original author(s) and source are credited.

Appendix 1: Comparison of bootstrap confidence intervals

We compare three types of bootstrap confidence intervals (c.i.):

  • simple percentile c.i.
  • bias-corrected percentile c.i.
  • accelerated bias-corrected percentile (BCa) c.i.

These c.i.’s are calculated as described in (Manly 1997), pp. 39–56. For the accelerated bias-corrected percentile c.i.’s calculation of the constant aacc is required. Manly (1997) suggests to approximate aacc by equation M203 with equation M204 the partial estimate of the parameter equation M205 based on all but the jth observation, and equation M206 the average of equation M207 (j = 1,…, N). In our case the parameter equation M208 is the fraction of common fragments pgs, estimated by either equation M209 or equation M210.

For equation M211 we take a pair of binary scores (y1j, y2j) (j = 1,…, N) to be an observation. The constant aacc is calculated by removing observation j from the pair of profiles, rescaling the fragment length distribution, calculating equation M212 from the reduced dataset, and repeating over all band positions (j = 1,…, N), resulting in partial estimates equation M213.

For equation M214 the information on band lengths is missing, and a pair of profiles can be summarized as a vector of counts (a, b, c, d). The observations are the pairs of binary scores 1–1 (occurring a times), 1–0 (b times), 0–1 (c times), and 0–0 (d times). The partial estimates equation M215 consist of weighted averages [with weights (a, b, c, d)] of equation M216 values. We label the weighted averages equation M217 (occurring a times), equation M218 (b times), equation M219 (c times), and equation M220 (d times). equation M221 is the weighted average of the 4 equation M222 values calculated for the profile pairs (a, b, c, d), (a − 1, b + 1, c, d), (a − 1, b, c + 1, d), (a − 1, b, c, d + 1), equation M223 is calculated from profile pairs (a + 1, b − 1,c, d), (a, b, c, d), (a, b − 1, c + 1, d), (a, b − 1, c, d + 1), equation M224 from profile pairs (a + 1, b, c  1, d), (a, b + 1, c − 1, d), (a, b, c, d), (a, b, c − 1, d + 1), and equation M225 from profile pairs (a + 1, b, c, d − 1), (a, b + 1, c, d − 1), (a, b, c + 1,d − 1), (a, b, c, d).

For the simulation dataset with 10,000 replicates, we calculated 95% bootstrap c.i.’s for equation M226, based on a bootstrap resample size of 1000. The results are shown in Table 5. The non-coverage rates for the 95% simple percentile c.i. range from 0.0497 to 0.0915 (average 0.0581), a bit larger than the nominal 0.05. The larger error rates occur for the profiles with smallest expected fragment counts (m = 40), and extreme values of pgs (pgs = 0.0, 0.9, 0.95). In general the c.i.’s are slightly too narrow. The 95% bias-corrected percentile c.i.’s have better non-coverage rates, ranging from 0.0475 to 0.0707 (average 0.0550). The non-coverage rates of the 95% BCa c.i.’s range from 0.0383 to 0.0545 (average 0.0486). This last method seems to be a bit too conservative, delivering intervals which are slightly too wide. Over the whole range of pgs values this last method performed best.

Table 5
Comparison of bootstrap confidence intervals for pgs from a simulation study on equation M143 for a range of values of genetic similarity pgs and expected numbers of fragments m1 and m2, 10,000 replicated pairs of AFLP profiles, 1,000 bootstrap resamples, fldFS from ...

For the same simulation data we calculated 95% bootstrap c.i.’s for equation M227 (see Table Table6).6). The non-coverage rates for the simple percentile method range from 0.0514 to 0.0878 (average 0.0584), for the bias-corrected method from 0.0499 to 0.065 (average 0.0548), and for the accelerated bias-corrected method from 0.0378 to 00555 (average 0.0487). Again, the accelerated bias-corrected method performs best with slightly conservative c.i.’s.

Table 6
Comparison of bootstrap confidence intervals for pgs from a simulation study on equation M145 for a range of values of genetic similarity pgs and expected numbers of fragments m1 and m2, 10,000 replicated pairs of AFLP profiles, 1,000 bootstrap resamples, fldFS from ...

Appendix 2

See Table 7 for a list of used symbols.

References


  • Althoff DM, Gitzendanner MA, Segraves KA (2007) The utility of amplified fragment length polymorphisms in phylogenetics: a comparison of homology within and between genomes. Syst Biol 56:477–484 [PubMed]

  • Bonin A, Ehrich D, Manel S (2007) Statistical analysis of amplified fragment length polymorphism data: a toolbox for molecular ecologists. Mol Ecol 16:3737–3758 [PubMed]

  • Caballero A, Quesada H, Rolán-Alvarez E (2008) Impact of amplified fragment length polymorphism size homoplasy on the estimation of population genetic diversity and the detection of selective loci. Genetics 179:539–554 [PMC free article] [PubMed]

  • Dice LR (1945) Measures of the amount of ecologic association between species. Ecology 26:297–302

  • DiCiccio TJ, Efron B (1996) Bootstrap confidence intervals. Stat Sci 11:189–228

  • Drossou A, Katsiotis A, Leggett JM, Loukas M, Tsakas S (2004) Genome and species relationships in genus Avena based on RAPD and AFLP molecular markers. Theor Appl Genet 109:48–54 [PubMed]

  • Duim B, Vandamme PAR, Rigter A, Laevens S, Dijkstra JR, Wagenaar JA (2001) Differentiation of Campylobacter species by AFLP fingerprinting. Microbiology 147:2729–2737 [PubMed]

  • Efron B, Tibshirani R (1993) An introduction to the bootstrap. Chapman & Hall, New York

  • Foulley JL, van Schriek MGM, Alderson L, Amigues Y, Bagga M, Boscher MY, Brugmans B, Cardellino R, Davoli R, Delgado JV, Fimland E, Gandini GC, Glodek P, Groenen MAM, Hammond K, Harlizius B, Heuven H, Joosten R, Martinez AM, Matassino D, Meyer JN, Peleman J, Ramos AM, Rattink AP, Russo V, Siggens KW, Vega-Pla JL, Ollivier L (2006) Genetic diversity analysis using lowly polymorphic dominant markers: the example of AFLP in pigs. J Hered 97:244–252 [PubMed]

  • Gort G, Koopman WJM, Stein A (2006) Fragment length distributions and collision probabilities for AFLP markers. Biometrics 62:1107–1115 [PubMed]

  • Gort G, Koopman WJM, Stein A, van Eeuwijk FA (2008) Collision probabilities for AFLP bands, with an application to simple measures of genetic similarity. JABES 13:177–198

  • Hansen M, Kraft T, Christiansson M, Nilsson N-O (1999) Evaluation of AFLP in Beta. Theor Appl Genet 98:845–852

  • Holland BR, Clarke AC, Meudth HM (2008) Optimizing automated AFLP scoring parameters to improve phylogenetic resolution. Syst Biol 57:347–366 [PubMed]

  • Jaccard P (1908) Nouvelles recherches sur la distribution florale. Bulletin de la Société Vaudoise des Sciences Naturelles. 44:223–270

  • Jansen J, van Hintum T (2007) Genetic distance sampling: a novel sampling method for obtaining core collections using genetic distances with an application to cultivated lettuce. Theor Appl Genet 114:421–428 [PubMed]

  • Koopman WJM, Gort G (2004) Significance tests and weighted values for AFLP similarities, based on Arabidopsis in silico AFLP fragment length distributions. Genetics 167:1915–1928 [PMC free article] [PubMed]

  • Koopman WJM, Zevenbergen M, Van den Berg R (2001) Species relationships in Lactuca s.l. (Lactuceae, Asteraceae) inferred from AFLP fingerprints. Am J Bot 88:1881–1887 [PubMed]
  • Koopman WJM, Hadam J, Doležel J (2002) Evolution of DNA content and base composition in Lactuca (Asteraceae) and related genera. In: Zooming in on the lettuce genome (Ph. D. thesis W.J.M. Koopman). Wageningen University, Wageningen, The Netherlands

  • Krauss SL (2000) Accurate gene diversity estimates from amplified fragment length polymorphism (AFLP) markers. Mol Ecol 9:1241–1245 [PubMed]

  • Manly BFJ (1997) Randomization, bootstrap and Monte Carlo methods in biology. Chapman & Hall, London

  • McCullagh P, Nelder JA (1991) Generalized linear models. Chapman & Hall, London

  • Mebrate SA, Dehne HW, Pillen K, Oerke EC (2006) Molecular diversity in Puccinia triticina isolates from Ethiopia and Germany. J Phytopathol 154:701–710

  • Meudt HM, Clarke AC (2007) Almost forgotten or latest practice? AFLP applications, analyses and advances. Trends Plant Sci 12:106–117 [PubMed]

  • Mueller UG, LaReesa Wolfenbarger L (1999) AFLP genotyping and fingerprinting. Trends Ecol Evol 14:389–394 [PubMed]

  • Nei M, Li WH (1979) Mathematical-model for studying genetic-variation in terms of restriction endonucleases. Proc Natl Acad Sci USA 76:5269–5273 [PMC free article] [PubMed]
  • R Development Core Team (2005) R: A language and environment for statistical computing. R Foundation for Statistical Computing. http://www.R-project.org, Vienna, Austria

  • O’Hanlon PC, Peakall R (2000) A Simple method for the detection of size homoplasy among amplified fragment length polymorphism fragments. Mol Ecol 9:815–816 [PubMed]

  • Reif JL, Melchinger AE, Frisch M (2005) Genetical and mathematical properties of similarity and dissimilarity coefficients applied in plant breeding and seed bank management. Crop Sci 45:1–7

  • Rieseberg LH (1996) Homology among RAPD fragments in interspecific comparisons. Mol Ecol 5:99–105
  • Robinson JP, Harris SA (1999) Amplified fragment length polymorphisms and microsatellites: a phylogenetic perspective. In: Which DNA marker for which purpose? Chap. 12. Final Compendium of the Research Project Development, optimization and validation of molecular tools for assessment of biodiversity in forest trees. European Union DGXII Biotechnology FW IV Research Programme Molecular Tools for Biodiversity. URL http://webdoc.sub.gwdg.de/ebook/y/1999/whichmarker/index.htm

  • Silvey SD (1975) Statistical inference. Chapman & Hall, London

  • Sneath PHA, Sokal RR (1973) Numerical taxonomy. Freeman, San Francisco

  • Tams SH, Melchinger AE, Bauer E (2005) Genetic similarity among European winter triticale elite germplasms assessed with AFLP and comparisons with SSR and pedigree data. Plant Breed 124:154–160

  • van Eeuwijk FA, Law JR (2004) Statistical aspects of essential derivation, with illustrations based on lettuce and barley. Euphytica 137:129–137

  • van Berloo R, Zhuy A, Ursem R, Verbakel H, Gort G, Van Eeuwijk FA (2008) Diversity and linkage disequilibrium analysis within a selected set of cultivated tomatoes. Theor Appl Genet 117:89–101 [PMC free article] [PubMed]

  • Vekemans X, Beauwens T, Lemaire M, Roldán-Ruiz I (2002) Data from amplified fragment length polymorphism (AFLP) markers show indication of size homoplasy and of a relationship between degree of homoplasy and fragment size. Mol Ecol 11:139–151 [PubMed]

  • Venzon DJ, Moolgavkar SH (1988) A method for computing profile-likelihood-based confidence intervals. Appl Stat 37:87–94

  • Vos P, Hogers R, Bleeker M, Reijans M, Vandelee T, Hornes M, Frijters A, Pot J, Peleman J, Kuiper M, Zabeau M (1995) AFLP: a new technique for DNA fingerprinting. Nucleic Acids Res 23:4407–4414 [PMC free article] [PubMed]

  • Wenzl P, Carling J, Kudrna D, Jaccoud D, Huttner E, Kleinhofs A, Kilian A (2004) Diversity arrays technology (DArT) for whole-genome profiling of barley. Proc Natl Acad Sci USA 101:9915–9920 [PMC free article] [PubMed]

  • Wong A, Forbes MR, Smith ML (2001) Characterization of AFLP markers in damselflies: prevalence of codominant markers and implications for population genetic applications. Genome 44:677–684 [PubMed]

  • Zhong D, Menge DM, Temu EA, Chen H, Yan G (2006) Amplified fragment length polymorphism mapping of quantitative trait loci for malaria parasite susceptibility in the yellow fever mosquito Aedes aegypti. Genetics 173:1337–1345 [PMC free article] [PubMed]

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