I am grateful to Dr. Weinberg for her insightful commentary (1) on my paper (2). Weinberg is looking forward eagerly to a day when researchers will be able to search the whole genome for susceptibility loci for complex diseases (1). Likewise, I notice with excitement that the conventional “risk factor” epidemiology as we know it has undergone a profound change. Moving into this postgenomic era, epidemiology can mean genemapping business, no less truly than it has been for so long about odds ratios and confidence intervals for, for example, smoking and lung cancer. To obtain a weighting scheme for the various family configurations encountered in actual practice, the Di is defined as the allele count for the case (Ci) minus the allele count for the following: 1) nontransmitted, Ni (case-parents data); 2) imputed nontransmitted, (casesibling data); 3) spouse, Si (case-spouse data); and 4) imputed spouse, (case-offspring data), where the “nontransmitted” refers to the parental alleles not transmitted to the case (1), the “imputed nontransmitted” has an allele count of , where the Bi is the mean allele count of the control siblings, and the spouse and the imputed spouse are defined the same as in my paper (2). The Dis defined in this way all have the same “mean” (e1 in my paper (2)) under the alternative hypothesis, irrespective of family configurations. The statistical efficiencies of the various family configurations are then in proportion to the inverses of their respective “variances” (e2 in my paper (2)). Table 1 presents the relative efficiencies for the various family configurations (relative to case-parents data). Note that, in this paper, the disequilibrium test (1), but not the transmission/disequilibrium test (2), was applied to the case-parents data. To be brief, I present only the case with the allele frequency set at P = 0.1. It is of interest to note the following. First, the case-spouse data and the case-parents data have exactly the same efficiency. Second, the case-offspring data and the casesibling data have roughly the same relative efficiency, when the number per family of offspring and the number per family of control siblings are equal. Third, the relative efficiency of the case-offspring (case-sibling) data with one offspring (control sibling) per family is ~0.5. These findings largely confirm Weinberg’s speculations (1). In addition, I agree with Weinberg (1) that the information from the following two lines of relatives can be integrated to improve the power of the disequilibrium test: line I, the parents or, if the parents are missing, the control siblings; and line II, the spouse or, if the spouse is missing, the offspring. The penultimate column of table 1 presents the relative efficiencies (also calculated from e2 in my paper (2)) for a Di, averaged from caseparents and case-spouse data, and the last column, the relative efficiencies from one control sibling and one offspring. It is of interest to note that the relative efficiencies are less than 2.00 for parents plus spouse. (Having two controls does not make the case twice as informative). Moreover, the relative efficiencies are less than 1.00 for one control sibling plus one offspring. (Two haploids do not make a whole diploid control). The values are larger than the corresponding values for two control siblings or two offspring, though. (Having two lines of relatives is better than having only one). The variance formula simplifies considerably under the null. We have, dropping the subscript i, Var(D) = Var(C – N) = Var(C) + Var(N) = 4P(1 – P), for caseparents data. Likewise, Var(D) = Var(C – S) = Var(C) + Var(S) = 4P(1 – P), for case-spouse data. Because the numbers of “identical by descent” (3) for a sibling pair are 2 (probability = 0.25), 1 (probability = 0.5), and 0 (probability = 0.25), we have Cov(B1, B2) = Cov(C, B1) = 0.25 × 2P(1 – P) + 0.5 × P(1 – P) + 0.25 × 0 = P(1 – P), with B1 and B2 being the allele count for the first and the second control siblings, respectively. Thus, for casesibling data with x control siblings per family,
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