bluuebunny/crossref_metadata_2025_split huggingface dataset:
No association of apolipoprotein A‐IV codon 347 and 360 variation with atherosclerosis and lipid transport in a sample of mixed hyperlipidemics
Authors: Carrejo Maurita H. , Sharrett A. Richey , Patsch Wolfgang , Boerwinkle Eric 
Published: June 2005

AbstractGenetic variation at the apolipoprotein (apo) A‐I/C‐III/A‐IV gene cluster on chromosome 11 has been associated with differences in occurrence of atherosclerosis and with variability in lipid levels among hypercholesterolemic‐hypertriglyceridemic individuals. The functional cause of the association is not known, but polymorphisms of the apo A‐IV gene are of interest because apo A‐IV is involved in both triglyceride and cholesterol metabolism. Two mutations in the apo A‐IV gene, 347T→S and 360Q→H, are known to cause amino acid substitutions in the mature protein. These polymorphisms were typed in a sample of 119 subjects with high cholesterol and high triglycerides in whom carotid artery wall thickness was previously shown to be strongly associated with silent polymorphic variation in the A‐I/C‐III/A‐IV gene cluster.The relative allele frequencies were 0.83 and 0.17 for codon 347T→S, and 0.95 and 0.05 for codon 360Q→H. These polymorphisms did not show a statistically significant relationship with prevalent hypertension, diabetes, or cardiovascular disease or with plasma lipid levels. Most importantly, these amino acid substitutions in apo A‐IV were not associated with carotid artery wall thickness. Therefore, the genetic cause of disease variability in a sample of mixed hyperlipidemics is not amino acid substitutions in codons 347 or 360 of the apolipoprotein A‐IV gene. ©1995 Wiley‐Liss, Inc.
Variation in HLA‐associated risks of childhood insulin‐dependent diabetes in the finnish population: II. Haplotype effects
Authors: Thomas Duncan , Pitkäniemi Janne , Langholz Bryan , Tuomilehto‐Wolf Eva , Tuomilehto Jaakko ,   DiMe Study Group
Published: June 2005

AbstractWe fitted models for the main effects of alleles at the HLA‐A, B, and DR loci and their haplotypes on the risk of insulin‐dependent diabetes mellitus (IDDM). Empirical Bayes methods were used, assuming independent exchangeable normal priors for effects at each locus separately and for haplotype effects. A pure main effects model, pure haplotype effects model, and a combined model were fitted using Gibbs sampling. The main effects model showed that the DR locus had the largest variation in risk between alleles, followed by the B locus; significance tests for each allele in this model were in general agreement with those in the companion paper [Langholz et al. (1995) Genet Epidemiol 12:441–453], although all relative risks were shrunk toward 1.0 in the empirical Bayes analysis. The variance estimate for pure haplotype effects was substantially larger than for any of the three main effects considered in this analysis, but in the combined model, the DR locus showed larger variability than the haplotype deviations. We confirmed that haplotype A2/B56/DR4 previously reported to be common in Finnish diabetics does indeed confer unusually high risk (relative risk = 7.6, P < 0.001), but found this to be only 1.9 times higher than predicted by its component main effects (P = 0.046). All of the other haplotypes could be adequately explained by their main effects. Empirical Bayes methods provide a natural means of dealing with the problems of multiple comparisons, multicolinearity, and sparse data that complicate the analysis of HLA data. ©1995 Wiley‐Liss, Inc.
Association of childhood rhabdomyosarcoma with neurofibromatosis type i and birth defects
Authors: Yang P. , Grufferman S. , Khoury M. J. , Schwartz A. G. , Kowalski J. , Ruymann F. B. , Maurer H. M. 
Published: June 2005

AbstractRhabdomyosarcoma (RMS) is an uncommon malignant soft tissue sarcoma whose cause is largely unknown. Reported risk factors include genetic alterations (e.g., p53 mutations, a defective gene at 11p15.5, or specific chromosomal translocation of t(2:13)), and parents' use of drugs around the time of conception. We present results from a national, case‐control study of 249 RMS cases (170 males and 79 females) and 302 controls (196 males and 106 females). The cases, aged 0–20 years at diagnosis, were identified via the Intergroup RMS Study‐III during 1982–1988. Controls were selected by random digit telephone dialing. As a supplement to the original study, information on genetic diseases and birth defects (BD) was collected from the subjects' parents by telephone interview. Fifty‐six (22.5%) cases and 55 (18.2%) controls were reported to have genetic diseases or BD (odds ratio [OR] = 1.30,95% confidence interval [CI] = 0.85–2.02, P = .21). The case group had a significantly higher frequency of neurofibromatosis type I (NF1) than did the control group, i.e., five cases (2.0%) had NF1 vs. zero controls (P = .02). The case group also had a higher frequency of major BDs than did the control group (6.0% vs. 2.6%, OR = 2.36, 95% CI = 0.92–6.52, P = .05). However, this excess was only observed in males (7.6% vs. 2.6%, OR = 3.16, 95% CI = 1.02–10.41, P = .02). Among the 15 cases having both RMS and major BDs, six (40.0%) had both conditions in the same regional anatomic site: Two (13.3%) had both in the extremities, two (13.3%) in the genitourinary system, and two in the head and neck. These findings suggest that common genetic mechanisms or in utero exposures may be invilved in the development of many childhood tumors and congenital abnormalities. ©1995 Wiley‐Liss, Inc.
Model‐free association analysis of a rare Disease
Authors: Bailey‐Wilson Joan E. , Sorant Benjamin , Sorant Alexa J. , Paul Christel M. , Elston Robert C. 
Published: June 2005

AbstractModel‐free methods of testing for association of a disease with alleles at a marker locus were used to analyze simulated data for a rare disease locus and 360 marker loci distributed at 2 cM intervals along six chromosomes. After adjustment for multiple tests, there was no evidence of significant heterogeneity of parental allele frequencies between a sample of parents with at least one affected child and a sample of parents with no affected children. Several significant deviations from Hardy‐Weinberg equilibrium were detected after Bonferroni correction but these were Type I errors. After adjusting for multiple tests, the model‐free test for association detected significant associations of alleles at two loci with the disease. These associations were in fact part of the generating model for the simulated data. However, these methods were unable to detect two other major loci contributing to the disease since these loci were not associated with any of the marker loci. ©1995 Wiley‐Liss, Inc.
Identification of susceptibility loci contributing to a complex disease using conventional segregation, linkage, and association methods
Authors: Falk C. T. , Ashley A. , Lamb N. , Sherman S. L. 
Published: June 2005

AbstractWe set out to apply conventional analytic methods to a GAW data set of nuclear families with an oligogenic disease that has a population prevalence of 0.023. We chose methods generally applied to disorders with at least one major gene. Our approaches included: (1) complex segregation analysis under two models of ascertainment, (2) linkage analysis assuming either a single‐locus trait with possible genetic heterogeneity or a two‐locus trait, and (3) allelic association studies using both a case/control approach and the haplotype relative risk (HRR) test. The association study was the only analysis of the three that provided evidence for genes playing a role in the etiology of this disorder. ©1995 Wiley‐Liss, Inc.
Effects of marker information on sib‐pair linkage analysis of a rare disease
Authors: Korczak Jeannette F. , Pugh Elizabeth W. , Premkumar Smita , Guo Xiuqing , Elston Robert C. , Bailey‐Wilson Joan E. 
Published: June 2005

AbstractModel‐free sib‐pair linkage analysis was used to screen the GAW9 ‐ Problem 1 data set for evidence of linkage of a rare disease to any of 360 highly polymorphic marker loci. Negative regressions nominally significant at the α = 0.05 level were obtained for 44 markers; however all of these proved to be Type I errors. None of the four disease loci were detected by sib‐pair linkage, which was not surprising, given the particular model and sampling scheme used to generate these data. Neither deleting parental marker genotypic information nor misspecifying marker allele frequency estimates substantially increased the Type I error rate. A two‐stage testing procedure using a 10 or 20 cM map and a liberal first stage significance level gave the same overall results as a one‐stage 2 cM map but required only about 42% or 22% as many markers, respectively. ©1995 Wiley‐Liss, Inc.
Integration of linkage analyses and disease association studies
Authors: Nemesure Barbara Berger , He Qimei , Mendell Nancy 
Published: June 2005

AbstractThe REGD procedure of the S.A.G.E. [1994] system was used to determine the mode of inheritance of the rare disease given in Problem 1. The likelihood ratio test statistic indicated that we should reject the hypotheses of dominant and recessive inheritance at the 0.01 level, so codominant inheritance was assumed. The estimated penetrance values computed from the β estimates given by the S.A.G.E. output were 1.0, 0.7, and 0.0 for the AA, AB, and BB genotypes respectively. A sample of three markers from each chromosome was used to determine which chromosome(s) gave evidence of having loci linked to the disease locus. The lod minus 0.83 support interval, which has been shown to provide the best approximation to 95% coverage among interval estimates [Nemesure et al., in press], was obtained for each of these markers. The criterion for rejecting the hypothesis of close linkage using the support interval methodology required that the left side of the lod minus 0.83 support interval about the maximum likelihood estimate, \documentclass{article}\pagestyle{empty}\begin{document}$ {\rm \hat \theta } $\end{document}, includes only values greater than θ = 0.10. This criterion suggested that chromosomes 2, 3, and 6 did not contain the disease genes. Classical lod‐score linkage analysis using the usual criteria of 3.0 for linkage and ‐2.0 for exclusion did not result in any regions being identified. On dropping the required lod score to 1.0, chromosomes 1, 3, and 6 gave results in favor of linkage with lod scores of 1.94 (θ = 0.19), 1.20 (θ = 0.24), and 1.30 (θ = 0.23), respectively. Association studies comparing unrelated cases to unrelated controls were done for all markers on all chromosomes. Two associations were observed which were significant at the 0.05 level after adjusting for the large number of multiple comparisons being made. The strongest association observed was between allele 7 of marker 23 on chromosome 5 and the disease (χ = 52.20, or = 4.7) and the second strongest was between allele 8 of marker 31 on chromosome 1 (χ = 20.10, OR = 3.4). ©1995 Wiley‐Liss, Inc.
TDT with covariates and genomic screens with mod scores: Their behavior on simulated data
Authors: Rice John P. , Neuman Rosalind J. , Hoshaw Stacy L. , Daw E. Warwick , Gu Chi 
Published: June 2005

AbstractWe describe an extension to the TDT (transmission/disequilibrium test) which allows for more than two marker alleles and for covariates measured on the parent or offspring. We also describe a systematic genomic search where the mod score (maximized lod score) is computed for each marker under constraints on the population prevalence or penetrances of a single locus. © 1995 Wiley‐Liss, Inc.
Genetic analysis of a common oligogenic trait with quantitative correlates: Summary of GAW9 results
Authors: Blangero J. 
Published: June 2005

AbstractThe statistical genetic analyses of a simulated data set for Genetic Analysis Workshop 9 Problem 2 are summarized. A diverse set of methods are evaluated with regard to the success with which they were able to determine the mode of inheritance and locate the genes underlying an oligogenically determined disease and four related quantitative traits. © 1995 Wiley‐Liss, Inc.
A brute force dichotomization approach to quantitative trait linkage analysis
Authors: Bailey Julia N. , Almasy Laura 
Published: June 2005

AbstractThe effect of dichotomizing a continuous phenotype in linkage analysis of a simulated oligogenic trait is explored. We conclude that dichotomization does not in itself preclude the detection of loci which account for as little as 16% of the genetic variance of the disease. The effects of inclusion of known covariates and quantitative trait linkage analysis are also discussed. © 1995 Wiley‐Liss, Inc.
Linkage analysis of a common oligogenic disease using selected sib pairs
Authors: Cardon Lon R. , Fulker David W. , Cherny Stacey S. 
Published: June 2005

AbstractSib pairs drawn from the simulated common oligogenic disease families were selected for extreme quantitative trait scores and analyzed using interval mapping and multipoint methods. Linkage analyses of 112 selected sib pairs, in which one or more members had trait values exceeding the disease threshold, were compared with analyses of the total unselected sib‐pair sample (771 pairs). Selected sample regression models yielded comparable significance levels to those obtained from the unselected sample at most loci on the six simulated chromosomes, demonstrating the efficiency of selected sib‐pair analysis for quantitative characters. Two of the three disease QTLs were detected in both selected and unselected samples. Interval mapping and multipoint analyses yielded location estimates close to the simulated positions of the QTLs. The combined strategy of using interval mapping and multipoint methods with selected sib pairs appears to provide improved accuracy and sensitivity over more traditional sib‐pair methods for detecting quantitative trait loci. © 1995 Wiley‐Liss, Inc.
Genetic epidemiologic analysis of quantitative phenotypes using gibbs sampling
Authors: Gauderman W. James , Witte John S. , Faucett Cheryl L. , Morrison John , Thomas Duncan C. 
Published: June 2005

AbstractWe analyzed two quantitative traits (Q1 and Q2) provided in the ‘Common Disease’ data set with the aim of detecting both genetic and environmental determinants. We used linear regression for screening measured variables, maximum likelihood segregation and linkage analyses for detecting and localizing unmeasured genes, and Gibbs sampling for joint segregation and linkage analyses with estimation of gene‐environment interaction and polygenic effects. For both Q1 and Q2, we successfully detected the unmeasured codominant major gene (MG) that was tightly linked to candidate gene C2. We also detected all of the measured variables used in generating Q1 (age, Q3, candidate gene C5) and Q2 (EF). Although our final models differed slightly from the true data generation models, our multifaceted analytic approach was successful in characterizing the determinants of Q1 and Q2. © 1995 Wiley‐Liss, Inc.
Use of exact and adjusted liability scores to detect genes affecting common traits
Authors: Goldin Lynn R. , Chase Gary A. , King Terri M. , Badner Judith A. , Gershon Elliot S. 
Published: June 2005

AbstractWe used the Haseman‐Elston sib‐pair test to screen for linkage of markers to genes for disease susceptibility in the simulated data as given in Problem 2 of GAW9. We applied the analysis to the underlying quantitative liability trait (Q1), other covariates of Q1 (Q2‐Q4), and the dichotomous affection status trait. In addition, we analyzed the residual Q1 after adjusting for the covariates. Using the sib‐pair linkage test, we identified a large region of chromosome 5 affecting the residual value of Q1, a region of chromosome 2 affecting Q3, and a region of chromosome 1 possibly affecting Q2. The analysis of the dichotomous traits did not reveal any regions to be significant. This is likely to be due to lack of information since the families were not selected through affected probands. © 1995 Wiley‐Liss, Inc.
A graphical approach for presenting linkage results from a genomic screen
Authors: Pugh Elizabeth W. , Mandal Diptasri M. , Wilson Alexander F. 
Published: June 2005

AbstractInitially, a sib‐pair linkage analysis was performed between the marker loci and six untransformed variables. Results from several variations of this initial analysis were compared using a graphical approach (P‐plots) to simplify presentation. When results were compared to the generating model, most of the aspects of the generating model were recovered, although we did not find evidence of the polygenic component shared by Q2 and Q3, or evidence of linkage between MG4 and Q4 at the 0.01 level. © 1995 Wiley‐Liss, Inc.
Analysis of familial breast cancer in genetic analysis workshop 9: Summary of findings
Authors: Goldgar David E. 
Published: June 2005

AbstractAs part of the 9th Genetic Analysis Workshop held in Val Morin, Quebec, October 16‐18, 1994, four workshop participants analyzed a large breast cancer data set. This data set consisted of phenotype and genetic marker data on 3884 individuals in 214 families with at least four cases of breast cancer contributed by members of an international breast cancer consortium. Two of the four papers [Barrett and Rigby; Commenges; this issue] utilized variants of affected pair methods to assess linkage of breast/ovarian cancer to the 17q markers in the data set. The third paper by Bansal et al. used a Monte‐Carlo approach to examine the question of intrafamilial clustering of breast and ovarian cancer. The last paper in the series [Leal and Ott] described a method of computing support intervals for risk estimates when there are uncertainties associated with the parameter estimates used to compute these risks. © 1995 Wiley‐Liss, Inc.
Development of the WPC approach and application to linkage analysis of breast cancer
Authors: Commenges Daniel 
Published: June 2005

AbstractWe develop the weighted pairwise correlation approach in several directions: we use a method based on martingale residuals rather than rank residuals; we give a measure of the strength of the linkage and we propose a method for multipoint analysis by combining the weights used in the statistics for different markers. These tools are applied to 214 breast cancer families. The values of the test statistics for the different markers are very similar to those obtained with the lod‐score approach; the multipoint analysis increases the value of the test statistic and is also in agreement with previous results. © 1995 Wiley‐Liss, Inc.
A Chromosome‐based method to infer IBD scores for missing and ambiguous markers
Authors: Gu Chi , Suarez Brian K. , Reich Theodore , Todorov Alexandre A. 
Published: June 2005

AbstractWe propose a probability model to impute missing identical‐by‐descent (IBD) vectors for linkage analysis, when adjacent marker loci are typed and interference is estimable. A chromosome‐based IBD distribution, conditioned on available marker data, is computed using a fast algorithm to estimate the joint probability of genes IBD at several equally spaced linked loci. Weighted IBD vectors are then used in various test statistics for linkage analysis. As an example, we analyzed the 18 affected sib pairs in the GAW9 Problem 1 data set using Risch's lod‐score test. © 1995 Wiley‐Liss, Inc.
The development of isolated and partially separated blastomeres of Amphioxus
Authors: Conklin Edwin G. 
Published: June 2005

AbstractBlastomeres of the egg of Amphioxus, separated along the plane of the first cleavage, develop into typical embryos and larvae because all materials of the egg are typically localized in each half. Blastomeres separated along the plane of the second cleavage never develop into typical larvae because the mesodermal crescent is largely lacking in the anterior cells and the chorda‐neural crescent lacking in the posterior ones. Isolated micromeres of the 8‐cell stage form blastulae with ectoderm but no endoderm, they never invaginate; macromeres form gastrulae wiht invaginated endoderm but never develop farther.Partially separated blastomeres develop into partially separated embryos and larvae. Anterior blastomeres of the 4‐cell stage separate more frequently than posterior ones and give rise to embryos and larvae double anteriorly but single posteriorly. Partially separated blastomeres often rotate with respect to each other, but each blastomere invariably preserves its original poles, axes, and germinal localizations. Such rotated blastomeres develop into double embryos and larvae with axes and poles as in the dislocated blastomeres; frequently antero‐posterior and dorso‐ventral axes are completely reversed in the twins.All axes and poles of the future larva are irreversibly determined at or before the first cleavage; complete regulation occurs only in the plane of bilateral symmetry; except for this bilateral regulation, development in Amphioxus, as also in Ascidians, is a mosaic work.
Fertilizing life of ferret sperm in the female tract
Authors: Chang M. C. 
Published: June 2005

AbstractFor the determination of the fertilizing life of ferret sperm in the female tract, gonadotrophin was intraperitoneally injected to induce ovulation at various times after the deposition of epididymal sperm into the uterine horns. The animals were examined 48 or 60 hours after ovulation injection to assess the proportion of fertilized eggs, although a few of them were allowed to deliver young.Intraperitoneal injection of 50 hypophysectomized R.U. of sheep pituitary extracts or 90 I.U. of Human Chorionic Gonadotrophin invariably induced ovulation in ferrets that were in heat. Ovulation occurred about 24 to 36 hours after the administration of H.C.G. and an average of 14.5 eggs with a range of 8 to 24 eggs per animal was obtained.The average number of sperm in the cauda epididymides and the vasa deferentia of nine males was 219 millions with a range of 24 millions at the end of breeding season to 621 millions at the peak of breeding season.The percentage of total eggs that were fertilized was 73, 40, 24, 15, 3, and 0 when the ovulation injection was given at 0, 36, 72, 84, 96, and 108 hours after sperm deposition respectively. Since ovulation occurred 30 hours after ovulation injection the fertilizing life of ferret sperm in the female tract is therefore no more than 126 hours. The percentage of eggs fertilized was lower following the deposition of epididymal sperm than following mating (73% vs. 98%). There is evidence that the fertility of sperm in the female tract was higher if a large number of sperm was deposited, and that the penetration of sperm into the egg facilitated the denudation of the egg.The majority of the penetrated eggs (54 to 84%) were at the pronuclear stage 48 hours after insemination and ovulation injection (i.e., about 18 hours after ovulation), at 60 hours (30 hours after ovulation) the majority of them were either at pronuclear stage (26–100%) or at the 2‐cell stage (46–57%), and at 70 hours (40 hours after ovulation) most of them were at the 4–6 cell stage. The morphology of fertilization in the ferret and the time sequence after sperm penetration were described.