Haplotypebased score test for linkage in nuclear families
 Chung Mo Nam^{1},
 Dae Ryong Kang^{1} and
 Jinheum Kim^{2}Email author
DOI: 10.1186/147121058277
© Nam et al; licensee BioMed Central Ltd. 2007
Received: 27 September 2006
Accepted: 01 August 2007
Published: 01 August 2007
Abstract
Background
To look for genetic linkage between angiotensinI converting enzyme(ACE) gene and hypertension in a Korean adolescent cohort, we developed a powerful test using the covariances between marginal differences and their variances in a transmission/nontransmission table.
Results
We estimated haplotype frequencies using the parental and affected offspring's genotypes and then constructed a transmission/nontransmission table for the parental haplotypes transmitted to the offspring. We then proposed a test for checking the marginal homogeneity in the table. Because the cells in the table were dependent due to the uncertainty of the parental haplotypes, we adopted a randomization procedure to estimate the significance of the observed test statistic. Simulations show that our test performs well on a nominal level and has a monotone power, which increases as the relative risk increases. With our test, there was no evidence of genetic linkage between the ACE gene and hypertension in the Korean adolescent cohort.
Conclusion
We developed a score test for linkage and used simulations to demonstrate that our test performs well at a nominal level. Under some situations where the diversity of haplotypes is low, the proposed test gained a little power over the method based on only variances between marginal differences in a transmission/nontransmission table.
Background
For linkage and/or association studies based on haplotypes, molecular haplotyping can be done for each individual, but at a high cost. Instead, statistical methods such as Clark's algorithm [1], the EM algorithm [2], or Gibb's sampler method [3], are commonly used to reconstruct haplotypes. The likelihood ratio test, which is based on the difference between the sum of the loglikelihoods of the case and control groups and the loglikelihood of the combined data, is usually used for casecontrol association studies [4], while the TDT is used for linkage and/or association studies in nuclear families. The latter compares the frequency of transmitted parental haplotypes with that of nontransmitted parental haplotypes to an affected offspring. One of the difficulties with the TDT is that there is uncertainty in estimating haplotype frequencies from parental genotypes. Wilson [5] and Clayton and Jones [6] proposed that these uncertain families be discarded from the analysis, but this drop in families leads to a low power. Clayton [7] proposed an estimating procedure based on likelihood, but it is no longer robust for population admixture or population stratification. Zhao et al [8] extended Spielman and Ewens' method [9] to test for marginal homogeneity in the transmission and nontransmission of haplotypes. Their method is advantageous over those of Wilson [5] and Clayton and Jones [6] because there is no discarding of unresolved families, and it is still robust for cases of population admixture or population stratification. Cordell and Clayton [10] modeled the nuclearfamily data via a conditional logistic regression in which they based on either haplotypes if parental phase may be inferrable or genotypes otherwise. Horvath et al [11] proposed familybased association tests with tightly linked markers where the phase may be ambiguous and the parental genotype data is missing. Their weighted conditional approach extended the method provided by Rabinowitz and Laird [12] to multiple markers.
Here, assuming that the haplotype block is very tight so that there is no recombination, we propose a score test for linkage and investigate its performance through simulations. We also illustrate the use of the proposed test with a data set taken from the Kangwha study [13].
Methods and Results
Score test
Let {H_{ s }H_{ u }, H_{ t }H_{ v }} denote the event in which the transmitted haplotype in the father is H_{ s }and the nontransmitted haplotype is H_{ t }, and simultaneously the transmitted haplotype in the mother is H_{ u }and the nontransmitted haplotype is H_{ v }. We designate {H_{ s }H_{ u }, H_{ t }H_{ v }} as one haplotype group and define {H_{ s }^{ g }H_{ u }^{ g }, H_{ t }^{ g }H_{ v }^{ g }} as haplotype groups corresponding to the set of genotypes g = 1, ..., G, where G is the number of distinct sets of genotypes across all markers. Here each set of genotypes refers to the observed genotypes for the individual markers of the two parents and the affected offspring. Suppose the total number of possible haplotypes is k.
where the summation over g ranges from 1 to G and the summations over s, t, u, and v range from 1 to k. They derived a test applying Spielman and Ewens' method to the table $\tilde{T}$.
Whenever all the parental haplotype phases are uniquely determined, T_{ s }asymptotically follows a chisquare distribution with (k  1) degrees of freedom because $\widehat{\Sigma}$ is the covariance matrix of $\widehat{\Delta}$ under no linkage (see Appendix for the details). We note, however, that the cells in $\widehat{T}$ are not independent because the contribution of a father or a mother is not limited to a cell because of its uncertainty as a haplotype pair, but reaches out to some cells according to the number of possible haplotype pairs and corresponding probabilities. That is why T_{ s }may not have a chisquare distribution with (k  1) degrees of freedom. Thus, we adopted the randomization test procedure introduced by Zhao et al [8] to estimate the statistical significance of the proposed test instead of relying on an asymptotic chisquare test.
Simulation studies
In our simulations, a hypothetical genealogy with three biallelic genetic markers was considered. We denoted alleles at the unobservable diseasesusceptibility locus as D and d. There are eight haplotypes consisting of bialleles at three loci, l_{1}, l_{2}, and l_{3}: H_{1} = (1,1,1), H_{2} = (1,1,2), H_{3} = (1,2,1), H_{4} = (1,2,2), H_{5} = (2,1,1), H_{6} = (2,1,2), H_{7} = (2,2,1), and H_{8} = (2,2,2). Here both H_{7} and H_{8} include the diseasesusceptibility allele d, whereas the other six haplotypes include the wildtype allele D. For each i = 1, ..., 8, let h_{ i }be the frequency of H_{ i }. We set 4 types of haplotype distribution: (h_{1}, h_{2}, h_{3}, h_{4}, h_{5}, h_{6}, h_{7}, h_{8}) = (.125, .125, .125, .125, .125, .125, .125, .125), (.343, .147, .147, .063, .147, .063, .027), (.343, .147, .210, .000, .210, .000, .063, .027), and (.700, .000, .000, .000, .210, .000, .063, .027), denoted as 'E', 'UE1', 'UE2', and 'UE3', respectively. Note that the degree of LD between each marker at loci, l_{1} and l_{2}, and the disease marker is high (D' = 1), while the LD between the marker at a locus l_{3} and the marker for the disease is relatively low (D' = 0 for E and UE1, D' = 0.15 for UE2, and D' = 1 for UE3). The subjects with the d allele have experienced mutations at loci, l_{1} and l_{2}, whereas the mutation at a locus l_{3} happened irrespective of the disease marker allele. Both dominant and recessive models were considered for the mode of inheritance. We assumed HardyWeinberg equilibrium and randommating and that the families were ascertained through one affected offspring.
Empirical type I error rates and powers of the proposed (T_{ s }) and Zhao et al's (T_{ z }) tests, and average percentages of perfectly estimated haplotypes for the dominant and recessive models according to 4 types of haplotype distribution (HD)
Dominant  Recessive  

HD  RR  p _{ f }  p _{ o }  T _{ s }  T _{ z }  p _{ f }  p _{ o }  T _{ s }  T _{ z } 
E  1  88  72  .060  .065  88  72  .045  .050 
2  88  73  .275  .280  88  72  .145  .110  
3  88  73  .635  .645  89  73  .275  .275  
4  89  74  .815  .820  89  72  .455  .460  
UE1  1  92  79  .060  .065  92  79  .050  .040 
2  92  79  .240  .235  92  79  .060  .050  
3  92  79  .560  .545  92  79  .050  .055  
4  92  79  .845  .840  93  79  .065  .055  
UE2  1  99  94  .050  .065  99  94  .075  .085 
2  98  92  .270  .250  99  93  .035  .040  
3  98  90  .610  .580  99  93  .065  .055  
4  97  89  .885  .860  99  93  .065  .060  
UE3  1  100  100  .040  .055  100  100  .090  .075 
2  100  100  .375  .325  100  100  .055  .065  
3  100  100  .795  .780  100  100  .060  .055  
4  100  100  .960  .950  100  100  .070  .060 
As mentioned by Rohde and Fuerst [15], the percentage of perfectly estimated haplotypes is higher when estimates are based on family trios rather than on parents alone. Therefore, using the offspring's genotype can reduce uncertainty in parental haplotypes. As expected, the lower the diversity of haplotypes, the higher the percentage of perfect haplotype estimates. For example, the percentage of perfectly estimated haplotypes for the UE1 case was higher than for the E case, and the percentage was higher for the UE2 or UE3 case than for the UE1 case. However, there was no difference in percentages between dominant and recessive models.
Based on our results with the proposed test, shown in entries corresponding to RR = 1, the estimated type I error rate seems to satisfy a nominal level of 0.05, regardless of the mode of inheritance and type of haplotype distribution. For the dominant models, the power of the test increases as RR increases from 2 to 4, irrespective of the type of haplotype distribution. In the recessive models, the power increases as RR increases for the E case, while the power equals to the nominal level for the UE1, UE2, and UE3 cases. This explains why the probability of a proband having the mutant homozygotes is very low at 0.006 for the UE1, UE2, and UE3 cases relative to 0.047 for the E case.
We also compared the proposed test and Zhao et al's test in terms of type I error rate and power. Both tests showed comparable performance at a significance level of 0.05. For the E case, Zhao et al's test was more powerful than our test; for the UE1, UE2, and UE3 cases, however, the latter had better power than the former. It seems that as the diversity of haplotypes decreases, the offdiagonal entries of $\widehat{T}$ correspondingly become more unbalanced and thereby are as informative as the marginal totals and this results in our test being more powerful than Zhao et al's test.
Empirical and asymptotic type I error rates of the proposed (T_{ s }) and Zhao et al's (T_{ z }) tests for the dominant model according to all the combinations of 4 pairs of haplotype distributions (HDs) and 3 types of proportion for the two populations in sample size (PP)
Empirical  Asymptotic  

Pair of HDs  PP  T _{ s }  T _{ z }  T _{ s }  T _{ z } 
E & E3  P11  .055  .045  .025  .020 
P13  .050  .045  .015  .030  
P31  .060  .035  .035  .040  
UE1 & UE3  P11  .060  .055  .025  .035 
P13  .055  .075  .020  .025  
P31  .050  .040  .030  .030  
UE2 & UE3  P11  .060  .060  .000  .015 
P13  .035  .030  .000  .005  
P31  .070  .055  .020  .035  
UE3 & UE3  P11  .045  .040  .005  .020 
P13  .030  .035  .000  .010  
P31  .050  .045  .005  .005 
Kangwha study
The Kangwha study was performed to analyze the natural history of BP in Koreans in order to determine important factors associated with changes in BP [13]. In 1986, we initially constructed a cohort of 430 6year old children who were living in Kangwha Province, Korea. The size of the cohort increased to 715 in 1992 and 784 in 1995. Our case group included blood samples from 101 students(61 boys and 40 girls) who experienced at least once systolic BP measurement of more than 130 mmHg or diastolic BP measurement of more than 85 mmHg between the ages of 15 and 17. Among the 101 samples from the case group, we were also able to obtain blood samples from the parents of 40 subjects and our analysis was based on these 40 probands and their parents.
Angiotensins are substances smaller than proteins that act as vasoconstricting agents, i.e., they cause blood vessels to narrow. Narrowing the diameter of blood vessels increases the blood pressure. ACE, which is located on chromosome 17q23 and has a length of 26 kb, converts angiotensin to its activated functional form, angiotensinII. There have been several studies investigating the relationship between the ACE gene and high BP [17]. In our analysis we included 4 SNPs, A240T and C93T on the promoter, I/D on intron 16, and A2350G on exon 17.
Observed values (pvalue) of proposed test (T_{ s }) and Zhao et al's test (T_{ z }) for assessing genetic linkage between ACE gene and hypertension
SNPs included  T _{ s }  T _{ z } 

all SNPs  5.275  7.363 
(A240T,C93T,I/D,A2350G)  (0.232)  (0.152) 
htSNPs  3.318  4.314 
(C93T,A2350G)  (0.332)  (0.163) 
In addition we examined whether any of the 4 SNPs were redundant and if exclusion of those SNPs could affect the outcomes of the two tests. We used a measure of PDE (proportion of diversity explained by a SNP set selection) proposed by Clayton [18] to select socalled htSNPs from the 4 SNPs being considered. This PDE acts like the coefficient of determination in an ordinary regression model. Using the HTSNP procedure [19] on 176 samples of the control group, we identified the htSNP set of C93T and A2350G with a PDE of 0.992. We repeated the two tests using only the two htSNPs and listed the corresponding results in the row denoted by 'htSNPs' in the first column of Table 3. As was in the case when using all SNPs, using just the htSNPs resulted in no evidence of genetic linkage between ACE and hypertension at a level of 0.05.
Discussion
Here we propose a score test for linkage between genetic markers and diseasesusceptibility genes based on haplotypes. First, we estimated haplotype frequencies using genotypes from affected offspring and their parents. As in [8], we constructed a transmission/nontransmission table of parents' haplotypes to the offspring. Then we proposed a test which mimics Sham's method [14] to test a marginal homogeneity in the transmission/nontransmission table.
Simulations indicate that our test works well at a nominal significant level. Further, we found that the power of the test is monotone and increases as the relative risk increases for dominant models. For recessive models with an unequal haplotype distribution, however, the test was highly conservative due to a low probability of a proband having the mutant homozygotes. In comparison with Zhao et al's test, their test had better power than our test for E case, while our test was more powerful than their test for the UE1, UE2, and UE3 cases. This implies our test has an advantage over Zhao et al's test in the situations where the diversity of haplotypes is low. We also found that our proposed test as well as Zhao et al's test is robust to population admixture. Although both asymptotic chisquared tests were conservative, Zhao et al's test seemed to be less conservative than our test.
Conclusion
We propose a score test for linkage between genetic markers and diseasesusceptibility genes and show through simulations that our test performed well at a nominal significant level. Our test showed the better performance in power than Zhao et al's test under some situations where the diversity of haplotypes is low. For an application to a Kangwha adolescent cohort, there was no remarkable evidence of genetic linkage between ACE and hypertension.
Appendix
If the haplotype phases for each parent were identified, the observed set of genotypes is compatible with only one haplotype group. Then ${\widehat{t}}_{ij}$ = t_{ ij }and thereby $\widehat{\Delta}$ = (t_{1}.  t._{1}, ..., t_{k1}.  t._{k1})', where t_{ ij }is the number of parents with haplotypes H_{ i }H_{ j }who transmit H_{ i }to the affected offspring. Thus the table $\widehat{T}$ = {t_{ ij }} can be simply considered as a transmission/nontransmission table corresponding to the case of a locus with k marker alleles because there are no ambiguities in the parental haplotypes. Let $n=2{\displaystyle {\sum}_{g=1}^{G}{n}_{g}}$. Define p_{ ij }as the probability of each parent with haplotypes H_{ i }H_{ j }transmitting H_{ i }to the affected offspring, and the marginal probabilities as ${p}_{i}.={\displaystyle {\sum}_{j=1}^{k}{p}_{ij}}$ and $p{.}_{j}={\displaystyle {\sum}_{i=1}^{k}{p}_{ij}}$. Following the arguments of Stuart [20], $\sqrt{n}[\widehat{\Delta}E(\widehat{\Delta})]$ asymptotically follows a (k  1)variate normal distribution with covariance n Σ = {nσ_{ ij }}, where E($\widehat{\Delta}$) = E(n(p_{1}.  p._{1}), ..., n(p_{k1}.  p._{k1}))' and σ_{ ii }= n[(p_{ i }. + p._{ i } 2p_{ ii })  (p_{ i }.  p._{ i })^{2}] and for i † j, σ_{ ij }= n[(p_{ ij }+ p_{ ji }) + (p_{ i }.  p._{ i })(p_{ j }.  p._{ j })]. From Sham and Curtis [21], under no linkage, p_{ i }. = p._{ i }, i = 1, ..., k. Therefore, to test linkage we can use the marginal homogeneity test statistic. Under no linkage, we have E($\widehat{\Delta}$) = (0, ..., 0)' and, plugging the maximum likelihood estimates, {t_{ ij }/n}, {t_{ i }./n}, and {t._{ j }/n}, of {p_{ ij }}, {p_{ i }.} and {p._{ j }}, respectively into the entries of Σ, the estimated covariance matrix n$\widehat{\Sigma}$ = {n${\widehat{\sigma}}_{ij}$}, where ${\widehat{\sigma}}_{ii}$ = t_{ i }. + t._{ i } 2t_{ ii }and for i ≠ j, ${\widehat{\sigma}}_{ij}$ = (t_{ ij }+ t_{ ji }). Hence T_{ s }asymptotically follows a chisquare distribution with (k  1) degrees of freedom.
Abbreviations
 ACE:

AngiotensinI converting enzyme
 BP:

Blood pressure
 EM:

ExpectationMaximization
 htSNP:

Haplotype tagging single nucleotide polymorphism
 RR:

Relative risk
 TDT:

Transmission/disequilibrium test
Declarations
Acknowledgements
This work was partially supported by the Korean Research Foundation Grant funded by the Korean Government (MOEHRD)(KRF2006312C00087)(J. Kim) and by a grant of the Korea Health 21 R&D Project, Ministry of Health and Welfare (03PJ1PG3210000015)(C.M. Nam).
Authors’ Affiliations
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