Eden R. Martin

Mitochondrial Polymorphisms Significantly Reduce the Risk of Parkinson Disease

Mitochondrial (mt) impairment, particularly within complex I of the electron transport system, has been implicated in the pathogenesis of Parkinson disease (PD). More than half of mitochondrially encoded polypeptides form part of the reduced nicotinamide adenine dinucleotide dehydrogenase (NADH) complex I enzyme. To test the hypothesis that mtDNA variation contributes to PD expression, we genotyped 10 single-nucleotide polymorphisms (SNPs) that define the European mtDNA haplogroups in 609 white patients with PD and 340 unaffected white control subjects. Overall, individuals classified as haplogroup J (odds ratio [OR] 0.55; 95% confidence interval [CI] 0.34-0.91; P=.02) or K (OR 0.52; 95% CI 0.30-0.90; P=.02) demonstrated a significant decrease in risk of PD versus individuals carrying the most common haplogroup, H. Furthermore, a specific SNP that defines these two haplogroups, 10398G, is strongly associated with this protective effect (OR 0.53; 95% CI 0.39-0.73; P=.0001). SNP 10398G causes a nonconservative amino acid change from threonine to alanine within the NADH dehydrogenase 3 (ND3) of complex I. After stratification by sex, this decrease in risk appeared stronger in women than in men (OR 0.43; 95% CI 0.27-0.71; P=.0009). In addition, SNP 9055A of ATP6 demonstrated a protective effect for women (OR 0.45; 95% CI 0.22-0.93; P=.03). Our results suggest that ND3 is an important factor in PD susceptibility among white individuals and could help explain the role of complex I in PD expression.

No Gene Is an Island: The Flip-Flop Phenomenon

An increasing number of publications are replicating a previously reported disease-marker association but with the risk allele reversed from the previous report. Do such flip-flop associations confirm or refute the previous association findings? We hypothesized that these associations may indeed be confirmations but that multilocus effects and variation in interlocus correlations contribute to this flip-flop phenomenon. We used theoretical modeling to demonstrate that flip-flop associations can occur when the investigated variant is correlated, through interactive effects or linkage disequilibrium, with a causal variant at another locus, and we show how these findings could explain previous reports of flip-flop associations.

Variation in the miRNA-433 Binding Site of FGF20 Confers Risk for Parkinson Disease by Overexpression of α-Synuclein

Parkinson disease (PD) is a common neurodegenerative disorder caused by environmental and genetic factors. We have previously shown linkage of PD to chromosome 8p. Subsequently, fibroblast growth factor 20 (FGF20) at 8p21.3-22 was identified as a risk factor in several association studies. To identify the risk-conferring polymorphism in FGF20, we performed genetic and functional analysis of single-nucleotide polymorphisms within the gene. In a sample of 729 nuclear families with 1089 affected and 1165 unaffected individuals, the strongest evidence of association came from rs12720208 in the 3 untranslated region of FGF20. We show in several functional assays that the risk allele for rs12720208 disrupts a binding site for microRNA-433, increasing translation of FGF20 in vitro and in vivo. In a cell-based system and in PD brains, this increase in translation of FGF20 is correlated with increased alpha-synuclein expression, which has previously been shown to cause PD through both overexpression and point mutations. We suggest a novel mechanism of action for PD risk in which the modulation of the susceptibility genes translation by common variations interfere with the regulation mechanisms of microRNA. We propose this is likely to be a common mechanism of genetic modulation of individual susceptibility to complex disease.

SNPing Away at Complex Diseases: Analysis of Single-Nucleotide Polymorphisms around APOE in Alzheimer Disease

There has been great interest in the prospects of using single-nucleotide polymorphisms (SNPs) in the search for complex disease genes, and several initiatives devoted to the identification and mapping of SNPs throughout the human genome are currently underway. However, actual data investigating the use of SNPs for identification of complex disease genes are scarce. To begin to look at issues surrounding the use of SNPs in complex disease studies, we have initiated a collaborative SNP mapping study around APOE, the well-established susceptibility gene for late-onset Alzheimer disease (AD). Sixty SNPs in a 1.5-Mb region surrounding APOE were genotyped in samples of unrelated cases of AD, in controls, and in families with AD. Standard tests were conducted to look for association of SNP alleles with AD, in cases and controls. We also used family-based association analyses, including recently developed methods to look for haplotype association. Evidence of association (P</=.05) was identified for 7 of 13 SNPs, including the APOE-4 polymorphism, spanning 40 kb on either side of APOE. As expected, very strong evidence for association with AD was seen for the APOE-4 polymorphism, as well as for two other SNPs that lie <16 kb from APOE. Haplotype analysis using family data increased significance over that seen in single-locus tests for some of the markers, and, for these data, improved localization of the gene. Our results demonstrate that associations can be detected at SNPs near a complex disease gene. We found that a high density of markers will be necessary in order to have a good chance of including SNPs with detectable levels of allelic association with the disease mutation, and statistical analysis based on haplotypes can provide additional information with respect to tests of significance and fine localization of complex disease genes.

HLA-DR2 Dose Effect on Susceptibility to Multiple Sclerosis and Influence on Disease Course

Models of disease susceptibility in multiple sclerosis (MS) often assume a dominant action for the HLA-DRB1*1501 allele and its associated haplotype (DRB1*1501-DQB1*0602 or DR2). A robust and phenotypically well-characterized MS data set was used to explore this model in more detail. A dose effect of HLA-DR2 haplotypes on MS susceptibility was revealed. This observation suggests that, in addition to the role of HLA-DR2 in MS, two copies of a susceptibility haplotype further increase disease risk. Second, we report that DR2 haplotypes modify disease expression. There is a paucity of benign MS and an increase of severe MS in individuals homozygous for DR2. Concepts of the molecular mechanisms that underlie linkage and association of the human leukocyte antigen (HLA) region to MS need to be revised to accommodate these data.

Analysis of European mitochondrial haplogroups with Alzheimer disease risk

We examined the association of mtDNA variation with Alzheimer disease (AD) risk in Caucasians (989 cases and 328 controls) testing the effect of individual haplogroups and single nucleotide polymorphisms (SNPs). Logistic regression analyses were used to assess risk of haplogroups and SNPs with AD in both main effects and interaction models. Males classified as haplogroup U showed an increase in risk (OR = 2.30; 95% CI, 1.03-5.11; P = 0.04) of AD relative to the most common haplogroup H, while females demonstrated a significant decrease in risk with haplogroup U (OR = 0.44 ; 95% CI, 0.24-0.80; P = 0.007). Our results were independent of APOE genotype, demonstrating that the effect of mt variation is not confounded by APOE4 carrier status. We suggest that variations within haplogroup U may be involved in AD expression in combination with environmental exposures or nuclear proteins other than APOE.

Parkin mutations and susceptibility alleles in late-onset Parkinson's disease

Parkin, an E2‐dependent ubiquitin protein ligase, carries pathogenic mutations in patients with autosomal recessive juvenile parkinsonism, but its role in the late‐onset form of Parkinsons disease (PD) is not firmly established. Previously, we detected linkage of idiopathic PD to the region on chromosome 6 containing the Parkin gene (D6S305, logarithm of odds score, 5.47) in families with at least one subject with age at onset (AAO) younger than 40 years. Mutation analysis of the Parkin gene in the 174 multiplex families from the genomic screen and 133 additional PD families identified mutations in 18% of early‐onset and 2% of late‐onset families (5% of total families screened). The AAO of patients with Parkin mutations ranged from 12 to 71 years. Excluding exon 7 mutations, the mean AAO of patients with Parkin mutations was 31.5 years. However, mutations in exon 7, the first RING finger (Cys253Trp, Arg256Cys, Arg275Trp, and Asp280Asn) were observed primarily in heterozygous PD patients with a much later AAO (mean AAO, 49.2 years) but were not found in controls in this study or several previous reports (920 chromosomes). These findings suggest that mutations in Parkin contribute to the common form of PD and that heterozygous mutations, especially those lying in exon 7, act as susceptibility alleles for late‐onset form of Parkinson disease. Ann Neurol 2003

Analysis of the RELN gene as a genetic risk factor for autism

Several genome-wide screens have indicated the presence of an autism susceptibility locus within the distal long arm of chromosome 7 (7q). Mapping at 7q22 within this region is the candidate gene reelin (RELN). RELN encodes a signaling protein that plays a pivotal role in the migration of several neuronal cell types and in the development of neural connections. Given these neurodevelopmental functions, recent reports that RELN influences genetic risk for autism are of significant interest. The total data set consists of 218 Caucasian families collected by our group, 85 Caucasian families collected by AGRE, and 68 Caucasian families collected at Tufts University were tested for genetic association of RELN variants to autism. Markers included five single-nucleotide polymorphisms (SNPs) and a repeat in the 5′-untranslated region (5′-UTR). Tests for association in Duke and AGRE families were also performed on four additional SNPs in the genes PSMC2 and ORC5L, which flank RELN. Family-based association analyses (PDT, Geno-PDT, and FBAT) were used to test for association of single-locus markers and multilocus haplotypes with autism. The most significant association identified from this combined data set was for the 5′-UTR repeat (PDT P-value=0.002). These analyses show the potential of RELN as an important contributor to genetic risk in autism.

Correcting for a Potential Bias in the Pedigree Disequilibrium Test

To the Editor: n nRecently, we proposed the pedigree disequilibrium test (PDT) as a test for allelic association and linkage (linkage disequilibrium) in general pedigrees (Martin et al. 2000). We have discovered that, in certain cases in extended pedigrees, the PDT can be biased under the null hypothesis. In this letter we describe the nature of the bias and illustrate a model in which the bias arises. We offer two alternative modifications to the PDT statistic, both of which result in valid tests of linkage disequilibrium over all genetic models and family structures. n n nIn constructing the PDT statistic, we considered two types of families within a pedigree. Informative nuclear families are those in which there is at least one affected individual, with both parents genotyped at the marker and at least one parent heterozygous. Informative discordant sibships have at least one affected and one unaffected sibling with different marker genotypes. For a marker locus with two alleles, M1 and M2, we defined the random variables XTj, for the jth triad (affected individual and both parents) in the pedigree, and XSj, for the jth discordant sib pair (DSP) in the pedigree: XTj = (no. of M1 transmitted) − (no. of M1 not transmitted) and XSj = (no. of M1 in affected sib) − (no. of M1 in unaffected sib), respectively. In our previous study (Martin et al. 2000), we defined a measure of association (D) for a pedigree containing nT triads from informative nuclear families and nS DSPs from informative discordant sibships: n n n n n nLet Di be the measure of association in the ith pedigree in a sample of N independent pedigrees. The PDT statistic is given by . The critical assumption is that the T is asymptotically normal, with mean 0 and variance 1, under the null hypothesis of no linkage disequilibrium. n n nThe difficulty that can be encountered is that, for some cases under the null hypothesis, the expected value of T may actually be different from 0, a situation that results in an inflated type I error. This is best illustrated by an example. Consider a fully penetrant dominant disease locus (with alleles d1 and d2) with no phenocopies, so that there is probability 1 that an individual with at least one copy of the disease allele is affected. Furthermore, assume that the disease allele (d1) is rare so that in any pedigree there is, at most, one segregating copy of the disease allele. Suppose that we have sampled extended three-generation pedigrees with the structure shown in figure 1. Only families in which the grandparent (GP2), parent (P2), and offspring (O) are all affected can lead to bias. Otherwise there will be, at most, one informative triad. Disease-locus genotypes are fully specified, given the affection status (see fig. 1). n n n nFigure 1 n nPedigrees illustrating PDT bias. Black circles denote affected individuals and white squares denote unaffected individuals. Disease- and marker-locus genotypes are shown for each individual. Values of quantities from equation (1) are given for each pedigree. ... n n n nSuppose that there is a marker locus fully linked to the disease locus (i.e., there is no recombination) but that there is no allelic association between the alleles at the two loci. This is the null hypothesis for the PDT. Furthermore, suppose that the marker locus has two alleles, with one allele—say, M1—being rare so that only one founder is heterozygous at the marker (families with no heterozygotes are not informative and therefore are not considered) and all three of the founders (GP1, GP2, and P1) are equally likely to be the heterozygote. The possible transmission patterns and the calculation of relevant quantities are shown in figure 1. For each pedigree we give the value of D, XT, and nT. Each of the six pedigrees in figure 1 is equally likely under the null hypothesis for the model given. Taking the expectation of D over these pedigrees yields E(D)=-xa01/6. Therefore, in this example, it is not the case that E(T)=0. n nFrom this example we can see that the problem arises when a grandparent is heterozygous: even though heterozygous grandparents are equally likely to transmit M1 or M2, the weights in the average (eq. [1]) differ depending on which allele is transmitted. If M1 is transmitted, then the average is over nT=2. If M2 is transmitted, then the average is over nT=1. Thus, there is a bias toward concluding that the more common allele is transmitted more often, even under the null hypothesis of equal transmission. n nOne can construct an unbiased test by requiring that the weights used in the average be independent of marker genotype. One alternative, hereafter referred to as the “PDT-avg,” is to average over all phenotypically informative units. Specifically, let nT be the number of fully genotyped family triads, irrespective of heterozygosity, and let nS be the number of DSPs, without requiring that they come from an informative sibship. For the example in figure 1, the PDT-avg is calculated by setting all nT=2, and this gives E(D)=0. A second alternative, hereafter referred to as the “PDT-sum,” is to use the sum from equation (1) and not use averages at all. This also gives E(D)=0 for the example in figure 1, since all nT=1 for the PDT-sum. (Software for the calculation of the PDT-avg and the PDT-sum statistics can be obtained from the Duke Center for Human Genetics Web site.) Approaches based on sums of random variables within pedigrees have also been proposed elsewhere (Martin et al. 1997; Teng and Risch 1999; Abecasis et al. 2000; Rabinowitz and Laird 2000). Intuitively, basing a statistic on the sum gives more weight to families with a greater number of phenotypically informative units, whereas averaging gives all families equal weight. n nTo compare these alternative tests to the original form of the PDT (“PDT-old”), we estimated type I error and power for the tests, using simulations (table 1). The same genetic models (i.e., 1–6) used by Martin et al. (2000) were used in these simulations, and marker- and disease-allele frequencies were set at .3. To simulate the null hypothesis, we simulated a lack of allelic association between the marker and disease loci but did not allow recombination. For each estimate, 5,000 replicate samples of 250 extended pedigrees of the structure used by Martin et al. (2000) were simulated. These simulations differ from those of Martin et al. (2000). In that study, ascertainment was assumed to be random with respect to affection status. In an attempt to simulate the ascertainment of extended pedigrees more realistically, the simulations in the present study produce pedigrees that are conditional on having at least one affected cousin pair. n n n nTable 1 n nEstimates of Type 1 Error and Power for PDTs for Various Genetic Models, Based on 5,000 Replicate Simulations[Note] n n n nThe results show that, for the cases examined, all tests have type I error levels close to the nominal level of .05 and thus are valid for these models. As we reported in our previous article (Martin et al. 2000), there is little bias reflected in the test when the original form of the statistic is used, although the estimates of type I error are larger than those for the PDT-avg and the PDT-sum, for each model. Part of the reason that there is little bias in these simulations is our choice of allele frequency for the marker. It can be shown that there is no bias when the frequencies of the marker alleles are .5. In these simulations, we used frequencies of .3 and .7, so there was less bias than expected for cases with more-extreme allele frequencies. An additional reason that little bias was seen is that we did not use grandparental genotypes in the calculation of the test statistic. It is possible to show that, when there are only two generations in the pedigree, there will be little bias as long as the genetic effect is not large. In these simulations, the genetic effect was low (i.e., the penetrances are all similar) for each of the models considered. It is noteworthy that, if grandparental genotypes were used, there could be bias even if there were no genetic effect due to this locus (i.e., even if penetrances were equal). n nThe results in table 1 demonstrate that the new tests can be more powerful than the original test. For these simulations we found that power is similar for the PDT-avg and the PDT-sum, but this will not always be the case. The PDT-sum gives more weight to families of larger size, whereas the PDT-avg gives all families equal weight. Thus, if pedigrees contain a similar number of phenotypically informative family units, then the values of the statistics will be similar. Differences will be most apparent when families are of different sizes. Exploratory simulations (not shown) have demonstrated that, in many cases, the PDT-sum can be more powerful than the PDT-avg; however, neither test is uniformly more powerful over all genetic models. n nIn summary, we have identified examples in which the original form of the PDT can be biased. There is no bias when the original form of the statistic is used in nuclear families (with or without parents) or when the marker-allele frequencies are .5. The bias is evident only when there are multiple generations contributing to the statistic, when the genetic effect due to the locus is strong, and when marker-allele frequencies are extreme. We have proposed two alternative statistics that not only remove the bias but also result in tests that can be more powerful than the original test. These tests provide valid alternatives for assessment of linkage disequilibrium in general pedigrees.

Analysis of linkage disequilibrium in γ-aminobutyric acid receptor subunit genes in autistic disorder

Autistic disorder (AD) is a neurodevelopmental disorder characterized by abnormalities in behavior, communication, and social interactions and functioning. Recently, Cook et al. reported significant linkage disequilibrium with an AD susceptibility locus and a marker, GABRB3 155CA-2, in the gamma-aminobutyric acid(A) (GABA(A)) receptor beta3-subunit gene on chromosome 15q11-q13. This linkage disequilibrium was detected using a multiallelic version of the transmission/disequilibrium test (TDT) in a sample of nuclear families having at least one child with autistic disorder. In an attempt to replicate this finding we tested for linkage disequilibrium with this marker, as well as with three additional markers in and around the GABA(A) receptor beta3-subunit gene, in an independent, clinically comparable set of AD families. Unlike Cook et al., we failed to detect significant linkage disequilibrium between GABRB3 155CA-2 and AD in our sample. We did, however, find suggestive evidence for linkage disequilibrium with a marker, GABRB3, approximately 60 kb beyond the 3 end of beta3-subunit gene. This finding lends support for previous reports implicating the involvement of genes in this region with AD. Am. J. Med. Genet. (Neuropsychiatr. Genet.) 96:43-48, 2000