A computational model for sex-specific genetic architecture of complex traits in humans: Implications for mapping pain sensitivity
- Chenguang Wang^{1},
- Yun Cheng^{1},
- Tian Liu^{1},
- Qin Li^{1},
- Roger B Fillingim^{2},
- Margaret R Wallace^{3},
- Roland Staud^{3},
- Lee Kaplan^{3} and
- Rongling Wu^{1}Email author
https://doi.org/10.1186/1744-8069-4-13
© Wang et al; licensee BioMed Central Ltd. 2008
Received: 29 November 2007
Accepted: 16 April 2008
Published: 16 April 2008
Abstract
Understanding differences in the genetic architecture of complex traits between the two sexes has significant implications for evolutionary studies and clinical diagnosis. However, our knowledge about sex-specific genetic architecture is limited largely because of a lack of analytical models that can detect and quantify the effects of sex on the complexity of quantitative genetic variation. Here, we derived a statistical model for mapping DNA sequence variants that contribute to sex-specific differences in allele frequencies, linkage disequilibria, and additive and dominance genetic effects due to haplotype diversity. This model allows a genome-wide search for functional haplotypes and the estimation and test of haplotype by sex interactions and sex-specific heritability. The model, validated by simulation studies, was used to detect sex-specific functional haplotypes that encode a pain sensitivity trait in humans. The model could have important implications for mapping complex trait genes and studying the detailed genetic architecture of sex-specific differences.
Background
Differences in males and females (sexual dimorphism) is ubiquitous in many biological aspects [1–3]. In humans, sexually dimorphic traits include those from morphological shapes and body size to brain development to disease susceptibility [4, 5]. Substantial differences are also observed in sensitivities to pain and pain-killing drugs, and susceptibility to developing chronic pain between men and women [6–8]. All these sex-specific differences are due to varying expression of genes on the X/Y chromosome and autosomes, thought to result from differences in cellular and hormonal environments between the two sexes [9]. A growing body of research has been conducted to elucidate the genetic control of sexual dimorphism in various complex phenotypes by gene mapping approaches [4, 5, 10, 11]. Despite these efforts, however, little is known about the genetic architecture underlying sex-related variation in a quantitative trait.
Since sex is easily determined, the effects of sex on morphological, developmental and pathological traits can be directly observed. However, characterizing the impacts of sex on the genetic architecture of these traits has been challenged by a lack of powerful statistical approaches. The motivation of this article is to develop a statistical and computational model that can systematically search for sex-specific genes contributing to quantitative variation and formulate testable hypotheses regarding the interplay between sexes and gene expression. Our model is principally different from those used in many previous studies that are aimed to detect sex-specific quantitative trait loci (QTLs) based on linkage or linkage disequilibrium analysis [2, 4, 5, 12, 13]. Our model will be founded on the statistical framework constructed by Liu et al. [14] to detect the effects and diversity of haplotypes constructed by single nucleotide polymorphisms (SNPs) that are genotyped at candidate genes or genome-wide [15]. Our model has been generalized to allow the test of sex differences in haplotype frequencies, allele frequencies and linkage disequilibria between different SNPs as well as additive and dominant effects of haplotypes on complex traits. It has power to identify sex-specific DNA sequence variants that encode complex phenotypes in men and women.
Model
Notation
The parameters contained in equation (1) can be used to describe some important aspects of the genetic structure and diversity of a natural population. Thus, differences in genetic architecture between the two different sexes can be characterized by these sex-specific parameters. Because it is easy to derive the closed forms for estimating haplotype frequencies [14], we will estimate the linkage disequilibrium from the estimated haplotype frequencies.
Diplotypes and their frequencies for each of nine genotypes at two SNPs, and composite diplotypes for one assumed sex-specific risk haplotype chosen from four possible haplotypes.
Diplotype | Risk Haplotype in Males | Risk Haplotype in Females | |||||||||
---|---|---|---|---|---|---|---|---|---|---|---|
Genotype | Configuration | Male Freq. | Female Freq. | [11] | [10] | [01] | [00] | [11] | [10] | [01] | [00] |
11/11 | [11] [11] | ${p}_{{\text{M}}_{11}}^{2}$ | ${p}_{{\text{F}}_{11}}^{2}$ | AA | $\overline{A}\overline{A}$ | $\overline{A}\overline{A}$ | $\overline{A}\overline{A}$ | BB | $\overline{B}\overline{B}$ | $\overline{B}\overline{B}$ | $\overline{B}\overline{B}$ |
11/10 | [11] [10] | $2{p}_{{\text{M}}_{11}}{p}_{{\text{M}}_{10}}$ | $2{p}_{{\text{F}}_{11}}{p}_{{\text{F}}_{10}}$ | $A\overline{A}$ | $A\overline{A}$ | $\overline{A}\overline{A}$ | $\overline{A}\overline{A}$ | $B\overline{B}$ | $B\overline{B}$ | $\overline{B}\overline{B}$ | $\overline{B}\overline{B}$ |
11/00 | [10] [10] | ${p}_{{\text{M}}_{10}}^{2}$ | ${p}_{{\text{F}}_{10}}^{2}$ | $\overline{A}\overline{A}$ | AA | $\overline{A}\overline{A}$ | $\overline{A}\overline{A}$ | $\overline{B}\overline{B}$ | BB | $\overline{B}\overline{B}$ | $\overline{B}\overline{B}$ |
10/11 | [11] [01] | $2{p}_{{\text{M}}_{11}}{p}_{{\text{M}}_{01}}$ | $2{p}_{{\text{F}}_{11}}{p}_{{\text{F}}_{01}}$ | $A\overline{A}$ | $\overline{A}\overline{A}$ | $A\overline{A}$ | $\overline{A}\overline{A}$ | $B\overline{B}$ | $\overline{B}\overline{B}$ | $B\overline{B}$ | $\overline{B}\overline{B}$ |
10/10 | $\{\begin{array}{c}[11][00]\\ [10][01]\end{array}$ | $\{\begin{array}{c}2{p}_{{\text{M}}_{11}}{p}_{{\text{M}}_{00}}\\ 2{p}_{{\text{M}}_{10}}{p}_{{\text{M}}_{01}}\end{array}$ | $\{\begin{array}{c}2{p}_{{\text{F}}_{11}}{p}_{{\text{F}}_{00}}\\ 2{p}_{{\text{F}}_{10}}{p}_{{\text{F}}_{01}}\end{array}$ | $\{\begin{array}{c}A\overline{A}\\ \overline{A}\overline{A}\end{array}$ | $\{\begin{array}{c}\overline{A}\overline{A}\\ A\overline{A}\end{array}$ | $\{\begin{array}{c}\overline{A}\overline{A}\\ A\overline{A}\end{array}$ | $\{\begin{array}{c}A\overline{A}\\ \overline{A}\overline{A}\end{array}$ | $\{\begin{array}{c}B\overline{B}\\ \overline{B}\overline{B}\end{array}$ | $\{\begin{array}{c}\overline{B}\overline{B}\\ B\overline{B}\end{array}$ | $\{\begin{array}{c}\overline{B}\overline{B}\\ B\overline{B}\end{array}$ | $\{\begin{array}{c}B\overline{B}\\ \overline{B}\overline{B}\end{array}$ |
10/00 | [10] [00] | $2{p}_{{\text{M}}_{10}}{p}_{{\text{M}}_{00}}$ | $2{p}_{{\text{F}}_{10}}{p}_{{\text{F}}_{00}}$ | $\overline{A}\overline{A}$ | $A\overline{A}$ | $\overline{A}\overline{A}$ | $A\overline{A}$ | $\overline{B}\overline{B}$ | $B\overline{B}$ | $\overline{B}\overline{B}$ | $B\overline{B}$ |
00/11 | [01] [01] | ${p}_{{M}_{01}}^{2}$ | ${p}_{{F}_{01}}^{2}$ | $\overline{A}\overline{A}$ | $\overline{A}\overline{A}$ | AA | $\overline{A}\overline{A}$ | $\overline{B}\overline{B}$ | $\overline{B}\overline{B}$ | BB | $\overline{B}\overline{B}$ |
00/10 | [01] [00] | $2{p}_{{\text{M}}_{01}}{p}_{{\text{M}}_{00}}$ | $2{p}_{{\text{F}}_{01}}{p}_{{\text{F}}_{00}}$ | $\overline{A}\overline{A}$ | $\overline{A}\overline{A}$ | $A\overline{A}$ | $A\overline{A}$ | $\overline{B}\overline{B}$ | $\overline{B}\overline{B}$ | $B\overline{B}$ | $B\overline{B}$ |
00/00 | [00] [00] | ${p}_{{M}_{00}}^{2}$ | ${p}_{{F}_{00}}^{2}$ | $\overline{A}\overline{A}$ | $\overline{A}\overline{A}$ | $\overline{A}\overline{A}$ | AA | $\overline{B}\overline{B}$ | $\overline{B}\overline{B}$ | $\overline{B}\overline{B}$ | BB |
If haplotypes triggers an effect on a quantitative trait, this means that at least one haplotype performs differently from the rest of the haplotypes. Without loss of generality, let haplotype [11] be such a distinct haplotype, called risk haplotype, designated as A. All the other non-risk haplotypes, [10], [01] and [00], are collectively expressed as $\overline{A}$. The risk and non-risk haplotypes form three composite diplotypes AA (symbolized as 2), $A\overline{A}$ (symbolized as 1) and $\overline{A}\overline{A}$ (symbolized as 0). The genotypic values of the three composite diplotypes may be different between the two sexes, arrayed in $({\mu}_{{\text{M}}_{2}},{\mu}_{{\text{M}}_{1}},{\mu}_{{\text{M}}_{0}})$ for the males and $({\mu}_{{\text{F}}_{2}},{\mu}_{{\text{F}}_{1}},{\mu}_{{\text{F}}_{0}})$ for the females, respectively. Let (a_{M}, d_{M}) and (a_{F}, d_{F}) be the additive and dominance genetic effects due to the risk and non-risk haplotypes in males and females, respectively.
Likelihoods
Assume that a total of n subjects (including n_{M} males and n_{F} females) sampled from the population are phenotyped for a quantitative trait. In each sex, there are nine possible genotypes for the two SNPs, each genotype with an observed number generally expressed as ${n}_{{r}_{1}{{r}^{\prime}}_{1}/{r}_{2}{{r}^{\prime}}_{2}}^{\text{M}}$ for the males and ${n}_{{r}_{1}{{r}^{\prime}}_{1}/{r}_{2}{{r}^{\prime}}_{2}}^{\text{F}}$ for the females (r_{1} ≥ r'_{1}, r_{2} ≥ r'_{2}, r_{3} ≥ r'_{3} = 1,0).
where (α_{M}, β_{M}) or (α_{F}, β_{F}) are body weight-related allometric coefficients, $({x}_{{\text{M}}_{i}},{z}_{{\text{M}}_{i}})$ or $({x}_{{\text{F}}_{i}},{z}_{{\text{F}}_{i}})$ are the indicator variables associated with the additive and dominance effects, respectively, and ${e}_{{\text{M}}_{i}}$ or ${e}_{{\text{F}}_{i}}$ is the residual error, normally distributed as $N(0,{\sigma}_{\text{M}}^{2})$ or $N(0,{\sigma}_{\text{F}}^{2})$. The genotypic values of composite diplotypes and variance are arrayed by a quantitative genetic parameter vector ${\Theta}_{{\text{M}}_{q}}=({\alpha}_{\text{M}},{\beta}_{\text{M}},{a}_{\text{M}},{d}_{\text{M}},{\sigma}_{\text{M}}^{2})$ for the males and ${\Theta}_{{\text{F}}_{q}}=({\alpha}_{\text{F}},{\beta}_{\text{F}},{a}_{\text{F}},{d}_{\text{F}},{\sigma}_{\text{F}}^{2})$ for the females, respectively.
is the relative proportion of diplotype [11] [00] within the double heterozygote for sex k.
It can be seen from equation (3) or (4) that maximizing $L({\Theta}_{{\text{M}}_{p}},{\Theta}_{{\text{M}}_{q}};{\Theta}_{{\text{F}}_{p}},{\Theta}_{{\text{F}}_{q}}|{y}_{\text{M}},{S}_{\text{M}},{y}_{\text{F}},{S}_{\text{F}})$ is equivalent to maximizing log $({\Theta}_{{k}_{p}}|{S}_{k})$ and log $L({\Theta}_{{k}_{p}}|{y}_{k},{S}_{k},{\Theta}_{{k}_{p}})$ individually in equation (5).
The EM algorithm
A closed-form solution for the EM algorithm [14] has been derived to estimate the unknown parameters that maximize the two sex-specific likelihoods of (5). The estimates of sex-specific haplotype frequencies are based on the log-likelihood function $({\Theta}_{{k}_{p}}|{S}_{k})$, whereas the estimates of sex-specific genotypic values of composite diplotypes and the residual variance are based on the log-likelihood function $L({\Theta}_{{k}_{p}}|{y}_{k},{S}_{k},{\widehat{\Theta}}_{{k}_{p}})$. These two different types of parameters can be estimated using a two-stage hierarchical EM algorithm (see ref. [14] for a detailed implementation).
Model selection
The largest likelihood value calculated is thought to correspond to the most likely risk haplotype. Under an optimal risk haplotype, we estimate sex-specific quantitative genetic parameters ${\widehat{\Theta}}_{{\text{M}}_{q}}$ and ${\widehat{\Theta}}_{{\text{F}}_{q}}$.
Hypothesis tests
The genetic architecture of a quantitative trait is characterized by population (including haplotype frequencies, allele frequencies, and linkage disequilibria) and quantitative genetic parameters (including haplotype effects and mode of inheritance for haplotypes). The model proposed provides a meaningful way for estimating the genetic architecture of a trait and further testing sex-specific differences in genetic control.
The influence of haplotypes on a quantitative trait is quantified in terms of the additive (a) and dominant genetic effects (d), and the mode of inheritance (ρ), which are estimated for each sex. Each of these population and quantitative genetic parameters can be tested when appropriate hypotheses are formulated.
Overall genetic control
where the L_{0} and L_{1} are the plug-in likelihood values under the null and alternative hypotheses of (8), respectively. Although the critical threshold for determining the existence of a functional haplotype can be based on empirical permutation tests, the LR may asymptotically follow a χ^{2} distribution with four degrees of freedom, so that the threshold can be obtained from the χ^{2}distribution table.
Sex-specific population genetic architecture
for the linkage disequilibrium between the two SNPs.
For each of the hypotheses (10)–(12), the LR values are calculated, which are each thought to asymptotically follow a χ^{2}-distribution with one degree of freedom. Sex-specific differences in overall population genetic architecture can be tested with the null hypothesis H_{0}: p_{M} = p_{F}, q_{M} = q_{F}, and D_{M} = D_{F}, with the corresponding LR value to be χ ^{2}-distributed with three degrees of freedom.
Sex-specific quantitative genetic architecture
The LR value calculated under the null and alternative hypotheses is suggested to follow a χ^{2}-distribution with two degrees of freedom. The rejection of the null hypothesis implies that the effects of the same haplotype are different between the two sexes. If there exists a sex-specific difference, the next step is to test whether this difference is due to the additive or dominant genetic effects, or both.
Sex-specific risk haplotypes
In the preceding sections, the same risk haplotype was assumed between the male and female populations. It is possible that the two sexes have different risk haplotypes. Let ${\mu}_{{\text{M}}_{{j}_{m}}}$ (j_{ m }= 2, 1, 0) and ${\mu}_{{\text{F}}_{{j}_{f}}}$ (j_{ f }= 2, 1, 0) be the genotypic values of composite diplotypes for the males and females constructed by a sex-specific rick haplotype. By reformulating the likelihood log $L({\Theta}_{{k}_{p}}|{y}_{k},{S}_{k},{\Theta}_{{k}_{p}})$ of equation (5) based on sex-specific composite diplotypes, these genotypic values can be estimated with the EM algorithm. A best combination of risk haplotypes between the two sexes can be determined from the AIC values.
Multi-locus haplotyping
Three-SNP model
for sex k.
Sex-specific population genetic architecture can be tested by comparing the differences in allele frequencies (p_{ k }, q_{ k }, r_{ k }) and linkage disequilibria of different orders $({D}_{{k}_{12}},{D}_{{k}_{13}},{D}_{{k}_{23}},{D}_{{k}_{123}})$ between the males and females.
In a natural population, there are 27 genotypes for the three SNPs. The frequency of each genotype is expressed in terms of haplotype frequencies. Some genotypes are consistent with diplotypes, whereas the others that are heterozygous at two or more SNPs are not. Each double heterozygote contains two different diplotypes. One triple heterozygote, i.e., 10/10/10, contains four different diplotypes, [111] [000] (in a probability of $2{p}_{{k}_{111}}{p}_{{k}_{000}}$), [110] [001] (in a probability of $2{p}_{{k}_{110}}{p}_{{k}_{001}}$), [101] [010] (in a probability of $2{p}_{{k}_{101}}{p}_{{k}_{010}}$) and [100] [011] (in a probability of 2p_{100}p_{011}). The relative frequencies of different diplotypes for this double or triple heterozygote are a function of haplotype frequencies (Supporting Information Table 1). The integrative EM algorithm can be employed to estimate the MLEs of haplotype frequencies. A general formula for estimating haplotype frequencies can be derived.
By assuming [111] as a risk haplotype (labeled by A) and all the others as non-risk haplotypes (labelled by $\overline{A}$), the formulation of genotypic values for three composite diplotypes, μ_{2} for AA, μ_{1} for $A\overline{A}$ and μ_{0} for $\overline{A}\overline{A}$ can be derived. Similar procedures described for the two-SNP model can be obtained to estimate and test sex-specific additive and dominance genetic effects when a haplotype contains three SNPs.
L-SNP model
It is possible that the two- and three-SNP models are too simple to characterize genetic variants for quantitative variation. We can develop a model that includes an arbitrary number of SNPs whose sequences are associated with the phenotypic variation. A key issue for the multi-SNP sequencing model is how to distinguish among 2^{ℓ-1} different diplotypes for the same genotype heterozygous at ℓ loci. The relative frequencies of these diplotypes can be expressed in terms of haplotype frequencies.
where D_{ k }'s are the linkage disequilibria of different orders among particular SNPs for sex k.
Sex-specific difference in terms of allele frequencies and linkage disequilibria between different SNPs as well as haplotype additive and dominance effects can be tested by formulating the corresponding hypotheses.
Results
Pain genetics study
The model proposed was used to detect differences in the genetic architecture of pain sensitivity between men and women. Genetic and phenotypic data were from a pain genetics project in which 237 subjects (including 143 men and 94 women) from five different races were sampled for six SNPs at three candidate genes. As a demonstration of the utilization of the model, we will focus on two SNPs, OPRDT80G (with two alleles T and G) and OPRDT921C (with two alleles T and C), at the delta opioid receptor. Pain testing procedures followed Fillingim et al. [16]. The phenotypic values of traits were subtracted by the means for each race to remove the effect due to races.
These two SNPs construct four haplotypes, [TC], [TT], [GC], and [GT], which yield 10 diplotypes, [TC] [TC], [TC] [TT], [TT] [TT], [TC] [GC], [TC] [GT], [TT] [GC], [TT] [GT], [GC] [GC], [GC] [GT], and [GT] [GT] and nine genotypes, TT/CC, TT/CT, TT/TT, TG/CC, TG/CT, TG/TT, GG/CC, GG/CT and GG/TT. Based on the observed numbers of each genotype in the male and female populations, we estimated sex-specific haplotype frequencies (Table 1). The pattern of haplotype distribution is consistent between the two sexes, with haplotypes [TC] and [TT] jointly occupying a majority proportion in the populations. Haplotype [GT] is very rare, with the frequency close to zero. SNP OPRDT80G has a low heterozygosity because the frequency of its commoner allele (T) is closer to 0.90, whereas there is a high heterozygosity for SNP OPRDT921C in terms of its averaged allele frequencies. The two SNPs are highly significantly associated at p = 3.41 × 10^{-5} for males and p = 9.63 × 10^{-5} for females, with a normalized linkage disequilibrium of D' = 1.00, because alleles T from OPRDT80G and T from OPRDT921C as well alleles G from OPRDT80G and C from OPRDT921C tend to form the same haplotypes more frequently than at random. There is no sex-specific difference in allele frequencies at the two SNPs and their linkage disequilibrium.
The estimates and tests of population genetic structure for two SNPs, OPRDT80G (with alleles T and G) and OPRDT921C (with two alleles C and T), and quantitative genetic effects of haplotypes constructed by these two SNPs on baseline pressure pain thresholds measured at the ulna in males and females.
Male | Female | |||||
---|---|---|---|---|---|---|
Genetic Parameter | MLE | p-value | MLE | p-value | Sex-specific LR | Sex-specific p-value |
Haplotype Frequency | ||||||
${\widehat{p}}_{TC}$ | 0.364 | 0.396 | ||||
${\widehat{p}}_{TT}$ | 0.505 | 0.463 | ||||
${\widehat{p}}_{GC}$ | 0.131 | 0.142 | ||||
${\widehat{p}}_{GT}$ | 0.000 | 0.000 | ||||
Allele Frequency and Linkage Disequilibrium | ||||||
${\widehat{p}}_{T}$ (OPRDT80G) | 0.869 | 0.858 | 0.079 | 0.778 | ||
${\widehat{q}}_{C}$ (OPRDT921C) | 0.495 | 0.537 | 0.578 | 0.447 | ||
$\widehat{D}$ | -0.066 3.41 × 10^{-5} | -0.066 9.63 × 10^{-5} | 0.001 | 0.973 | ||
Additive and Dominant Effects and Inheritance Mode for Risk Haplotype [TT] | ||||||
$\widehat{a}$ and $\widehat{d}$ | 0.129 | 0.548 | 31.44 | 1.49 × 10^{-7} | ||
$\widehat{a}$ | -0.2 | 0.431 | 0.3 | 0.481 | 4.72 | 0.030 |
$\widehat{d}$ | 0.7 | 0.058 | -0.5 | 0.350 | 24.86 | 6.16 × 10^{-7} |
$\widehat{\rho}$ | 2.9 | 1.8 | 0.63 | 0.427 |
Monte carlo simulation
Simulation studies were performed to test the statistical properties of the model proposed. Given a certain sample size (n), we simulated two SNPs by assuming different allele frequencies and linkage disequilibria between two sexes. The hypothesized allele frequencies at the two SNPs are p_{M} = 0.5, q_{M} = 0.6 and D_{M} = 0.1 for males and p_{F} = 0.8, q_{F} = 0.9 and D_{F} = 0.06 for females. By postulating one of the four haplotypes constructed by the two SNPs as a risk haplotype, we calculated the genetic variance among three composite diplotypes using the additive effect a = 0.6 and dominance effect d = 0.8, from which the residual variance was calculated when the heritability (H^{2}) of a trait is given. The phenotypic values of the trait were simulated by assuming that they follow a normal distribution under four different simulation designs, (1) n = 100 and H^{2} = 0.1, (2) n = 400 and H^{2} = 0.1, (3) n = 100 and H^{2} = 0.4, and (4) n = 400 and H^{2} = 0.4.
The MLEs of population and quantitative genetic parameters and the standard errors of the estimates obtained by the new model and the power to detect sex-specific differences under different simulation designs. Parameter estimates by a conventional model are also given.
Sex | |||||
---|---|---|---|---|---|
Simulation Design | Genetic Parameter | Male | Female | Power | Traditional Model |
1 | p | 0.5025 ± 0.0360 | 0.7829 ± 0.0265 | 0.95 | 0.6433 ± 0.0224 |
q | 0.5964 ± 0.0349 | 0.8732 ± 0.0193 | 0.95 | 0.7347 ± 0.0201 | |
D | 0.0961 ± 0.0194 | 0.0181 ± 0.0099 | 0.80 | 0.0779 ± 0.0120 | |
a | 0.6055 ± 0.2008 | 0.3045 ± 0.1277 | 0.24 | 0.4789 ± 0.1192 | |
d | 0.3061 ± 0.3371 | 0.6093 ± 0.1798 | 0.19 | 0.5843 ± 0.1875 | |
2 | p | 0.5013 ± 0.0172 | 0.7980 ± 0.0141 | 1.00 | 0.6496 ± 0.0111 |
q | 0.6011 ± 0.0177 | 0.8958 ± 0.0101 | 1.00 | 0.7484 ± 0.0100 | |
D | 0.0986 ± 0.0098 | 0.0084 ± 0.0053 | 1.00 | 0.0775 ± 0.0061 | |
a | 0.6050 ± 0.1114 | 0.3056 ± 0.0728 | 0.67 | 0.4907 ± 0.0616 | |
d | 0.2980 ± 0.1604 | 0.5986 ± 0.0933 | 0.42 | 0.5715 ± 0.0966 | |
3 | p | 0.5008 ± 0.0366 | 0.7835 ± 0.0269 | 1.00 | 0.6421 ± 0.0221 |
q | 0.5945 ± 0.0338 | 0.8731 ± 0.0195 | 1.00 | 0.7338 ± 0.0194 | |
D | 0.0965 ± 0.0178 | 0.0161 ± 0.0088 | 0.86 | 0.0747 ± 0.0115 | |
a | 0.5989 ± 0.0924 | 0.3012 ± 0.0505 | 0.85 | 0.4582 ± 0.0483 | |
d | 0.2934 ± 0.1352 | 0.6021 ± 0.0652 | 0.64 | 0.5355 ± 0.0715 | |
4 | p | 0.5014 ± 0.0181 | 0.7974 ± 0.0136 | 1.00 | 0.6494 ± 0.0113 |
q | 0.6000 ± 0.0171 | 0.8968 ± 0.0102 | 1.00 | 0.7484 ± 0.0099 | |
D | 0.1001 ± 0.0092 | 0.0077 ± 0.0048 | 1.00 | 0.0745 ± 0.0058 | |
a | 0.6003 ± 0.0439 | 0.3010 ± 0.0301 | 1.00 | 0.4684 ± 0.0251 | |
d | 0.3008 ± 0.0659 | 0.6006 ± 0.0372 | 0.99 | 0.5499 ± 0.0360 |
The estimation of quantitative genetic parameters including the additive (a) and dominance effects (d) needs the determination of an optimal risk haplotype. When all possible risk haplotypes were assumed for the simulated data, we found that the true risk haplotype gave the largest likelihood among the four possible cases. In general, quantitative genetic parameters can well be estimated, but the estimation precision increases dramatically with sample size and heritability. Although the additive effect can be obtained with reasonable precision at a modest heritability (0.1) with a modest sample size (100), the precise estimation of the dominance effect relies upon a larger heritability and sample size. Also, the given difference in the additive effect between two sexes can be detected with great power, even when both the sample size and heritability are small. But the same size of sex-specific difference in the dominance effect can be detected with the same power only when the sample size is 400 and heritability is 0.4 (Table 3). The traditional model gave biased estimates of the sex-specific additive and dominance effects regardless of increasing sample size and heritability.
We conducted an additional simulation study, in which the data simulated under the assumption of no sex-specific differences in all genetic parameters were analyzed by the new and traditional model. As expected, both the models provide reasonable estimates of population and quantitative genetic parameters, with estimation precision increasing with increasing sample size and heritability (data not shown). This, in conjunction with the results in Table 3, suggests that the new model provides a general tool for study the genetic architecture of a complex trait, regardless of whether the genetic control of the trait is sex-specific.
Discussion
The genetic architecture of a quantitative trait is complex in terms of interactions between its underlying genetic factors and various environments including sex [1–3]. However, in many current studies, gene by sex interactions are often ignored simply because existing analytical models are not incorporated by environmental factors. While different phenotypic expressions of a trait between the two sexes can be easily measured [4, 5], sex-specific discrepancy in the genetic control of the trait can be discerned only when a sophisticated model is used. There is strong evidence for sex-specific genetic influence [17, 18] even for traits that display no sexual dimorphism [19]. Although quantitative genetic models are available to estimate sex-specific heritabilities due to aggregative effects of many genes [5, 19–22] or map sex-specific QTLs for phenotypic variation [4, 5, 11–13, 23], the model proposed in this article can dissect sex-specific genetic control at the DNA sequence levels.
Our model is founded on the conceptual framework for haplotyping a trait with single nucleotide polymorphisms (SNPs) formulated by Liu et al. [14]. Since haplotypes constructed by physically associated SNPs are thought to affect the expressivity of a complex trait [23, 24], it is more precise to characterize such haplotype effects by incorporating gene by gene and gene by sex interactions. Lin and Wu [25] extended Liu et al.'s [14] model to estimate haplotype-haplotype interactions. The new model reported here can not only estimate sex-specific genetic parameters, but also provide a series of statistical procedures for testing sex-specific differences in the genetic architecture of quantitative variation. In a natural population, the structure and pattern of genetic variation can be studied by population genetic parameters, such as haplotype frequencies, allele frequencies and linkage disequilibria. Thus, the understanding of differences in these parameters between the two sexes help to infer the sex-specific genetic structure of a natural population and its evolutionary processes. As shown through simulation studies, our model is alert to discern sex-specific differences in basic population genetic parameters.
Liu et al. [14] assumed a so-called reference or risk haplotype that triggers an effect on complex traits in a different way from the other haplotypes. Thus, the combinations between the risk and non-risk haplotypes (composite diplotypes) will perform differently, depending on the type of combination, i.e., risk by risk, risk by non-risk and non-risk by non-risk. Liu et al. [14] proposed the concepts of the additive effect due to the substitution of the non-risk (or risk) haplotype by the risk (or non-risk) haplotype and the dominance effect due to the interaction between the risk and non-risk haplotypes. These concepts have been integrated into the current model that allows the test of additive by sex and dominance by sex interaction effects. Simulation studies suggest that the new model displays adequate power to detect differences in these quantitative genetic parameters between the sexes, although the detection of sex-specific dominance effects needs a much larger sample size and/or heritability level.
Our model was used to analyze a real data set for pain genetics. Our analyses of a pain sensitivity trait-baseline pressure pain threshold measured at the ulna-by estimating genetic parameters and testing their sex-specific differences in a combination of male and female samples revealed that males and females have different population structure at two SNPs genotyped from a candidate gene (delta opioid receptor) for human pain and that haplotypes exert different genetic effects on the trait between the two sexes. A further test indicates that the risk haplotype [TT] detected by the model exemplifies sex-specific modes of inheritance in affecting the pain trait. While there are no differences among composite diplotypes in males, a significant additive effect was detected in females. Both additive and dominance effects due to the risk haplotype identified are different between the two sexes. Anholt and Mackay [2] described three major mechanisms that explain sex-specific difference in trait control, i.e., sex-specific effects (a gene affects only one sex), sex-biased effects (a gene affects both sexes but to different degrees), and sex-antagonistic effects (a gene affects both sexes but in opposite directions). In our example, the functional haplotype detected affects the pain trait in a sex-antagonistic effect manner, a mechanism thought to help the maintenance of genetic variation in natural populations [26].
In practice, failure to model sex-specific architecture may significantly hamper the ability to detect signals of functional genetic variants in genomewide screens. Although combining male and female data to increase sample size are tempting approaches to increase power, the estimates in this way will be biased from true sex-specific differences. Possible mechanisms that cause sex differences include parent-of-origin effects [27], linkage to or interaction with sex chromosomes, or differences arising from sex-specific hormonal environments. Our gene by sex interaction model that is incorporated by these mechanisms can be modified to consider interactions between genes and any other environments such as life style. Our interaction models should provide a more powerful tool to draw a detailed and precise picture of the genetic architecture of any complex traits that are important to human health.
Declarations
Acknowledgements
The preparation of this manuscript is partially supported by a joint NSF/NIH grant (0540745) to RW and an NIH grant (R01 NS041670-05A1) to RBF.
Authors’ Affiliations
References
- Mackay TF: The genetic architecture of quantitative traits: lessons from Drosophila . Curr Opin Genet Dev 2004, 14(3):253–257. 10.1016/j.gde.2004.04.003View ArticlePubMedGoogle Scholar
- Anholt RR, Mackay TFC: Quantitative genetic analyses of complex behaviours in Drosophila . Nature Reviews Genetics 2004, 5: 838–849. 10.1038/nrg1472View ArticlePubMedGoogle Scholar
- Foerster K, Coulson T, Sheldon BC, Pemberton JM, Clutton-Brock TH, Kruuk LEB: Sexually antagonistic genetic variation for fitness in red deer. Nature 2007, 447: 1107–1110. 10.1038/nature05912View ArticlePubMedGoogle Scholar
- Weiss LA, Pan L, Abney M, Ober C: The sex-specific genetic architecture of quantitative traits in humans. Nature Genetics 2006, 38: 218–222. 10.1038/ng1726View ArticlePubMedGoogle Scholar
- Weiss LA, Abney M, Cook EH, Ober C: Sex-specific genetic architecture of whole blood serotonin levels. American Journal of Human Genetics 2005, 76: 33–41. 10.1086/426697PubMed CentralView ArticlePubMedGoogle Scholar
- Fillingim RB: Sex-related influences on pain: a review of mechanisms and clinical implications. Rehabilitation Psychology 2003, 48: 165–174. 10.1037/0090-5550.48.3.165View ArticleGoogle Scholar
- Fillingim RB: Sex, gender and pain: The biopsychosocial model in action XX vs. XY. The International Journal of Sex Differences in the Study of Health, Disease and Aging 2003, 1: 98–101.Google Scholar
- Craft RM, Mogil JS, Aloisi AM: Sex differences in pain and analgesia: the role of gonadal hormones. European Journal of Pain 2004, 8: 397–411. 10.1016/j.ejpain.2004.01.003View ArticlePubMedGoogle Scholar
- Rinn JL, Snyder M: Sexual dimorphisms in mammalian gene expression. Trends in Genetics 2005, 21: 298–305. 10.1016/j.tig.2005.03.005View ArticlePubMedGoogle Scholar
- Mogil JS, Richards SP, O'Toole LA, Helms ML, Mitchell SR, Kest B, Belknap JK: Identification of a sex-specific quantitative trait locus mediating nonopioid stress-induced analgesia in female mice. Journal of Neuroscience 1997, 17: 7995–8002.PubMedGoogle Scholar
- Zhao W, Ma CX, Cheverud JM, Wu RL: A unifying statistical model for QTL mapping of genotype-sex interaction for developmental trajectories. Physiological Genomics 2004, 19: 218–227. 10.1152/physiolgenomics.00129.2004View ArticlePubMedGoogle Scholar
- Butterfield RJ, Roper RR, Rhein DM, Melvold RW, Haynes L, Ma RZ, Doerge RW, Teuscher C: Sex-specific QTL govern susceptibility to Theilers murine en-cephalomyelitis virus-induced demyelination (TMEVD). Genetics 2003, 163: 1041–1046.PubMed CentralPubMedGoogle Scholar
- Femandez JR, Vogler GP, Tarantino LM, Vignetti S, Plomin R, McClearn GE: Sex-exclusive quantitative trait loci influences in alcohol-related phenotypes. American Journal of Medical Genetics 1999, 88: 647–652. 10.1002/(SICI)1096-8628(19991215)88:6<647::AID-AJMG13>3.0.CO;2-6View ArticleGoogle Scholar
- Liu T, Johnson JA, Casella G, Wu RL: Sequencing complex diseases with HapMap. Genetics 2004, 168: 503–511. 10.1534/genetics.104.029603PubMed CentralView ArticlePubMedGoogle Scholar
- The International HapMap Consortium: The International HapMap Project. Nature 2003, 426: 789–794. 10.1038/nature02168View ArticleGoogle Scholar
- Fillingim RB, Kaplan L, Staud R, Ness TJ, Glover TL, Campbell CM, Mogil JS, Wallace MR: The A118G single nucleotide polymorphism of the mu-opioid receptor gene (OPRM1) is associated with pressure pain sensitivity in humans. Journal of Pain 2005, 6: 159–167. 10.1016/j.jpain.2004.11.008View ArticlePubMedGoogle Scholar
- Jensen H, Sather B-E, Ringsby TH, Tufto J, Griffth SC, Ellegren H: Sexual variation in heritability and genetic correlations of morphological traits in house sparrow ( Passer domesticus ). Journal of Evolutionary Biology 2003, 16: 1296–1307. 10.1046/j.1420-9101.2003.00614.xView ArticlePubMedGoogle Scholar
- Leips J, Mackay TFC: Quantitative trait loci for life span in Drosophila melanogaster: interactions with genetic background and larval density. Genetics 2002, 155: 1773–1788.Google Scholar
- Schousboe K, Visscher PM, Henriksen JE, Hopper JL, Sorensen TIA, Kyvik KO: Twin study of genetic and environmental influences on glucose tolerance and indices of insulin sensitivity and secretion. Diabetologia 2003, 46: 1276–1283. 10.1007/s00125-003-1165-xView ArticlePubMedGoogle Scholar
- van Beijsterveldt CE, Bartels M, Hudziak JJ, Boomsma DI: Causes of stability of aggression from early childhood to adolescence: a longitudinal genetic analysis in Dutch twins. Behavioral Genetics 2003, 33: 591–605. 10.1023/A:1025735002864View ArticleGoogle Scholar
- Keski-Rahkonen A, Neale BM, Bulik CM, Pietil KH, Pietilainen KH, Rose RJ, Kaprio J, Rissanen A: -bf Intentional weight loss in young adults: Sex-specific genetic and environmental effects. Obesity Research 2005, 13: 745–753. 10.1038/oby.2005.84View ArticlePubMedGoogle Scholar
- Ober C, Pan L, Phillips N, Parry R, Kurina LM: Sex-specific genetic architecture of asthma-associated quantitative trait loci in a founder population. Curr Allergy Asthma Rep 2006, 6(3):241–246. 10.1007/s11882-006-0041-4View ArticlePubMedGoogle Scholar
- Bader JS: The relative power of SNPs and haplotype as genetic markers for association tests. Pharmacogenomics 2001, 2(1):11–24. 10.1517/14622416.2.1.11View ArticlePubMedGoogle Scholar
- Judson R, Stephens JC, Windemuth A: The predictive power of haplotypes in clinical response. Pharmacogenomics 2000, 1: 5–26. 10.1517/14622416.1.1.15View ArticleGoogle Scholar
- Lin M, Wu RL: Detecting sequence-sequence interactions for complex diseases. Current Genomics 2006, 7: 59–72. 10.2174/138920206776389775View ArticleGoogle Scholar
- Rice WR: Sexually antagonistic genes: experimental evidence. Science 1992, 256: 1436–1439. 10.1126/science.1604317View ArticlePubMedGoogle Scholar
- Reik W, Walter J: Genomic imprinting: parental influence on the genome. Nature Reviews Genetics 2001, 2: 21–32. 10.1038/35047554View ArticlePubMedGoogle Scholar
Copyright
This article is published under license to BioMed Central Ltd. This is an Open Access article distributed under the terms of the Creative Commons Attribution License (http://creativecommons.org/licenses/by/2.0), which permits unrestricted use, distribution, and reproduction in any medium, provided the original work is properly cited.