Objective: The aim of this study was to investigate variations in the endothelial cell protein C receptor gene (EPCRgene) that may play a role in thrombosis and the effects of these variations on the plasma soluble endothelial cell proteinC receptor (sEPCR) level in Turkish patients with venous thrombosis.Material and Methods: This study included 111 thrombosis patients and 73 healthy controls. Following DNAextraction, PCR, SSCP, and DNA sequencing analysis of 4 exons of the EPCR gene was performed. Plasma sEPCR wasmeasured via enzyme-linked immunosorbent assay (ELISA).Results: In all, 3 polymorphisms were detected in exons 1-4. C3998T (SNP no: rs2069952) polymorphism was detectedin intron 2 and C4678G (A1 haplotype) (SNP no: rs9574) in the 3’ untranslated region (3’UTR). There weren’t anysignificant differences in C3998T polymorphism between the control and patient groups. There wasn’t a significantdifference in plasma sEPCR levels between both controls and patients that carried the A1 allele. A4600G substitution (A3haplotype) (SNP no: rs867186) was observed in exon 4 and was associated with a 2.04-fold higher risk of thrombosis.A3 allele carriers had higher sEPCR levels than those without the allele. Mean sEPCR level in the patients with thehomozygous A3 allele was 289 ng μL–1, versus 113.42 ng μL–1 in those with the homozygous A1 allele.Conclusion: The A1 haplotype might offer protection against thrombosis and the A3 haplotype might be associatedboth with elevated plasma sEPCR and elevated risk of venous thrombosis in the Turkish population. Plasma sEPCRlevels were significantly higher in those that carried the A3 allele (4600A>G) (both patients and controls). Among theparticipants that carried the A1 allele (4678C>G), plasma sEPCR did not differ significantly. The protein C (PC) anticoagulant pathway is physiologicallyimportant to the regulation of coagulation.While the PC pathway plays a crucial role in controllingthrombosis, it also has anti-inflammatory functions -3.The esThe PC pathway is a natural anticoagulation mechanismthat prevents excessive thrombin generation. WhenThrombin binds to TM, PC is activated about 1000 timesfaster than free thrombin . PC actiEPCR binds to PC and also APC with high affinityand promotes PC activation on endothelium by increasincreasingthe catalytic efficiency of the thrombin-TM complex-6. AlthoPrimary defects of the PC pathway increase the riskof venous thrombosis . A metalPlasma sEPCR is elevated in patients with systemicinflammatory diseases . sEPCR bThe human EPCR gene is located on chromosome20q11.2 and is composed of 4 exons and 3 introns. Exon1 encodes the 5’ untranslated region (UTR) and the signalpeptide, exons 2 and 3 encode most of the extracellularregion of EPCR, and exon 4 encodes an additional10 residues of the extracellular region of EPCR, the transmembranedomain, the cytoplasmic tail, and the 3’ UTR,17. In aTo date, more than 92 polymorphisms and 5 diseaserelatedmutations have been observed on the human EPCR gene ,20. AmonThe study included 111 patients and 73 controls.The patient group included 64 females (57.6%) and 47males (42.4%), and the control group included 48 females(65.7%) and 25 males (n34.3%). Mean age in the controlgroup was 28.5 years, versus 26.01 years in the patientgroup. None of the control group participants had a personaland a family history of venous thrombosis. All participantspresented to Ankara University, School of Medicine,Department of Pediatric Molecular Genetics. All theparticipants provided written informed consent and thestudy was approved by the Ankara University School ofMedicine Ethics Committee.Plasma sEPCR was measured via enzyme-linked immunosorbentassay (ELISA)  . DNA wasSingle strand conformation polymorphism (SSCP) wasperformed to determine if there was band pattern variationamong the participants. SSCP was performed in 143 of theparticipants to detect the polymorphisms. Sequencing ofthe EPCR gene was performed when SSCP yielded a differentband pattern . Direct sequencing of exon 4 was performedto detect A4600G mutation, whereas sequencing and REanalysis were used to detect C4678G mutation. All theparticipants were genotyped using 4678G to C substitution,which creates a Ddel restriction site (5’C/TNAG 3’).The samples were incubated using Ddel restriction enzyme at 37 °C for 16 h, andthen the digestion products were electrophoresed on 3%agarose gel .Statistical analysis was performed using TADPOLEv.2.01 for Windows. The level of statistical significancewas set at P <0.05. The odds ratio (OR) and 95% confidenceinterval (CI) were calculated based on the logisticmodel. Allele frequencies were calculated via gene counting.In all these analyses the group with homozygous normal(4678 CC) served as the reference category to whichrisk was expressed.SSCP band pattern variation was noted in exons 3 andthe first region of exon 4. Three SNPs in exons3 and 4 that have been previously described; (C3998T)rs2069952 ,32,33, lNumerous variations in the EPCR gene have been reported. Any defect in the EPCR gene that leads to are duction in receptor expression or impaired receptor function may stimulate the development of thrombosis.A 23-bp duplication of nucleotides between 4189 and4213 at the genomic level, according to NT_011362.10(a number of previous publications has referred this as position 4031) , resultsA polymorphism in exon 4 (A3 haplotype) encodes serine instead of glycine at codon position 219 (NM 006404.3);this occurs in the trans membrane region of EPCR .This hapNavarro et al. reported that mean age at initial onset of thrombosis was higher in non-carriers of the rs867186(A4600G) allele (44 ± 14 years) than in carriers (35 ± 8years), and that the probability of not having thrombosis at age 40 years was lower in prothromb in 20210A carriers with the EPCR A4600G allele. They also reported that the presence of the A4600G allele, plasma sEPCR >147 ngmL–1, and prothrombin >129% all were associated withan increase in the risk of thrombosis. The risk of venousthromboembolism is influenced by the prothrombin level and the EPCR A3 haplotype, due to their effect on sEPCR levels . SmokingThe results of the present study confirm that there isa strong correlation between the A3 haplotype and elevatedplasma sEPCR, as previously reported ,23,29.ThThe authors of this paper have no conflicts of interest,including specific financial interests, relationships, and/or affiliations relevant to the subject matter or materialsincluded.This study was supported by the Turkish Society of Hematology.


MicroRNAs (miRNAs) are small RNA molecules that regulate the expression of corresponding messenger RNAs (mRNAs). Variations in the level of expression of distinct miRNAs have been observed in the genesis, progression and prognosis of multiple human malignancies. The present study was aimed to investigate the association between four highly studied miRNA polymorphisms  and cancer risk by using a two-sided meta-analytic approach.An updated meta-analysis based on 53 independent case-control studies consisting of 27573 cancer cases and 34791 controls was performed. Odds ratio (OR) and 95% confidence interval (95% CI) were used to investigate the strength of the association.P = 0.004, TT vs. CC) while other miRNA SNPs showed no association with overall cancer risk. Subgroup analyses based on type of cancer and ethnicity were also performed, and results indicated that there was a strong association between miR-146a rs2910164 and overall cancer risk in Caucasian population under recessive model . Stratified analysis by cancer type also associated mir-196a2 rs11614913 with lung and colorectal cancer at allelic and genotypic level.Overall, the pooled analysis showed that mir-196a2 rs11614913 was associated with a decreased cancer risk (OR = 0.846, The present meta-analysis suggests an important role of mir-196a2 rs11614913 polymorphism with overall cancer risk especially in Asian population. Further studies with large sample size are needed to evaluate and confirm this association. MicroRNAs (miRNAs) are a class of endogenous, small nonprotein-coding single-stranded RNA molecules of ∼22 nucleotides in length that regulate a broad range of biologic and pathologic processes Cancer is eventually an outcome of chaotic expression of genes involved in developmental, cell growth and differentiation processes. Recent studies have implicated miRNAs in the genesis, progression  and prognosis of multiple human malignancies Single nucleotide polymorphisms (SNPs) are the most common form of variation present in the human genome. SNPs present in the miRNA gene regions can alter their expression and/or maturation leading to aberrant miRNA regulation. Many epidemiological studies have examined the association of SNPs in microRNAs with cancer susceptibility . HoweverWe searched the PubMed, Medline and Embase databases using the search terms “miRNA,” “cancer/carcinoma,” and “polymorphism/variant” updated until August 25, 2012 and limited to English language papers. Identification of meta-analyses of association studies on miRNA polymorphisms and cancer was also carried out through a search of electronic databases of PubMed, Medline and Embase, up to August 2012. The Medical Subject Headings and key words used for the search were “miRNA”, “cancer”, “polymorphism”, and “meta-analysis” . The online searching was accompanied by checking reference lists from the identified articles and reviews for potentially eligible original reports.All miRNA association studies were included in the present meta-analysis if they met the following criteria: 1) case-control study, 2) outcome cancer , and 3) sufficient data for examining an odds ratio (OR) with 95% confidence interval (95% CI). The major exclusion criteria were as follows: 1) duplicate data, 2) case reports, series, abstract, comment, review and editorial and 3) insufficient data. Articles published in a language other than English were also excluded.From each study, information like: author, year of publication, country of origin, cancer type, ethnicity, number of cases and controls, source of control groups (study design) and genotyping method was extracted. In some cases, identical data were described in more than one publication; in such cases the secondary studies were not included in the meta-analysis. In a few studies, part of the data had already been reported elsewhere, therefore, only the novel data was included. We also checked for HWE in control subjects among all publications.Genotype distributions were extracted from the eligible publications for each polymorphism or computed from allele frequencies (if genotype frequencies were not reported) on the basis of sample size, assuming Hardy-Weinberg equilibrium (HWE).The quality of selected studies was evaluated by scoring according to a set of predetermined criteria. The categories in scoring system used for assessing study quality are summarized in 2-based Q-test among the studies 2 test with one degree of freedom was performed in controls to observe deviation from HWE. Publication bias was weighted by Begg’s funnel plot and Egger’s linear regression method with P<0.05 being considered statistically significant In the present meta-analysis, we investigated the potential association between the variant allele of miRNA polymorphisms and cancer risk. Also, analysis between the heterozygote, the homozygote and also in dominant and recessive models was done to estimate cancer risk. Stratified analyses were performed by tumor site, ethnicity and source of controls . Other potentially relevant sub-group analyses such as age, sex and cancer subgroup could not reliably be investigated due to limited data availability. Between-study heterogeneity was evaluated with a χFor evaluating impact of HWE-deviated studies on point estimates in genotype based contrasts, ORs were corrected by using the HWE-predicted genotype count in controls instead of the observed counts, as recommended by Trikalinos et al. P-value = 0.320; CC vs. GG; Q = 77.126, P = <0.001, 2I = 66.289%, CC vs. GG). Similar results were obtained at the allelic level and for other genetic models also with significant heterogeneity . Removing 8 low score studies {The miR-146a rs2910164 polymorphism was analyzed in 27 studies with 12088 cases and 17340 controls and was not found to be associated with cancer risk . However, this association was lost in Asian populations . Removing low scoring studies did not make significant deviation from the above obtained results {Thirty studies investigated mir-196a2 rs11614913 polymorphism and its association with cancer (13703 cases and 15439 controls). The “T” allele of the polymorphism was considered as the variant allele in the present analysis. The overall OR showed statistically significant association between the rs11614913 polymorphism and reduced risk of cancer  . HoweverIn the stratified analysis by ethnicity, Asian individuals had lower risk of cancer under both the allelic and genotypic level (OR<1.0), whereas Caucasian individuals did not show any significant association under any genetic model. Additional subgroup analysis significantly associated rs11614913 ‘T’ allele with decreased cancer risk in population based study design (OR = 0.77–0.90) .P = 0.046, I2 = 72.85%  and OR = 1.177, 95%CI = 1.007–1.377, P = 0.041, I2 = 74.66% (CT vs. TT); P = 0.066; allelic model). However, no association was found between genotype CC and cancer risk under the other models. Based on the ethnicity of study population, association was found in Asian populations under allelic and recessive models  .The effect of some polymorphisms could not be evaluated due to the limited number of studies .A single study involved in the meta-analysis was removed each time to reflect the influence of the individual data set to the pooled ORs for each of the studied miRNA polymorphisms. The corresponding pooled ORs were not significantly altered for any of the SNPs studied -d.p = 0.398 for allelic model) (P2tailed = 0.646). Similarly for mir-196a2 rs11614913 and mir-499 rs3746444, funnel plots were symmetrical and the Egger's test for both models showed no significance, suggesting little evidence of publication bias  = −3.047 to 1.253; t = 0.859, c model) . Also, Bion bias  and S3.A cumulative meta-analysis was also done by sorting the studies in the sequence of largest to smallest, and analysis performed with the addition of each study. The point estimate of the study did not deviate with the addition of smaller studies, ruling out the possibility of publication bias for all the analyzed miRNA SNPs.Eleven meta-analyses published in 2011 and 2012 were retrieved, focusing on 2 miRNA polymorphisms (miR-146a rs2910164 and miR-196a2 rs11614913). We also computed the population-attributable risk (PAR) to refer to the proportion of disease risk in Caucasians and Asians that can be attributed to the causal effects of the risk SNP (variant genotype). PAR can be assessed by using the formula In the present study, we reviewed the available literature on genetic studies of miRNA SNPs in cancer and conducted four independent meta-analyses for association between overall cancer and mir-146a rs2910164, mir-196a2 rs11614913, mir-149 rs2292832 and mir-499 rs3746444 polymorphisms. Our results associated mir-196a2 rs11614913 with a decreased overall cancer risk. Meanwhile, there was no association between other studied miRNA SNPs. However, due to the small number of studies with an overall mediocre quality and lack of confirmatory studies, it is very difficult to draw any definitive conclusions.miR-146a, first found in mouse, has been shown to play an important role in tumorigenesis by promoting cell proliferation and colony formation in NIH/3T3 cells Aberrant mir-196a2 expression is implicated in cancer susceptibility and metastasis in several malignancies The mir-499 microRNA has also been implicated in several human malignancies . A T>C , howeverFor mir-149 rs2292832, no statistical association was found in the overall comparison and subgroup analysis. Because of the limited number of studies (7) for this polymorphism, the results should be interpreted with caution.For other miRNA polymorphisms, because of limited number of studies (ranging from one to three), meta-analysis was not done as it would not have been reliable.One of the important concerns in every meta-analysis is publication bias. Because meta-analysis reviews quantitative data from numerous studies, the publication bias effect of the literature incorporated in the study can bias the meta-analytic outcome. In the present study, the funnel plot for overall results was symmetrical for all the analyzed miRNA SNPs, indicating negligible likelihood of publication bias. The Egger’s test and Begg and Mazumdar rank correlation test were also negative for publication bias. However, the possibility of publication bias cannot completely be ruled out To the best of our knowledge, the present study is the most comprehensive meta-analysis to date to have assessed the relationship between the miRNA polymorphisms and cancer risk. Nevertheless, our meta-analysis had some limitations common to these types of studies. First, the present meta-analysis only included case-control studies, most of which were hospital based and excluded 12 cohort studies to avoid potential heterogeneity in comparing results. Thus, the controls may not reflect the representative element of the source population. Second, the difference in the geographic areas  and genetic backgrounds of the study cohort in each article could influence the results. Third, the low sample size in some of the included studies might influence the statistical power to better evaluate the association between miRNA polymorphisms and overall cancer, especially in subgroup analysis. Fourth, gene-gene and gene-environment interactions were not analyzed which might alter the associations between miRNA gene polymorphisms and cancer. Also, a more precise analysis stratified by variables such as age, sex etc. could not be performed due to limitations of the data which also restricted our ability to detect possible sources of heterogeneity.In conclusion, the results of our meta-analysis demonstrate that mir-196a2 rs11614913 polymorphisms have significant associations with overall cancer risk, although some results are limited by the small number of studies. However, no significant association exists between mir-146a rs2910164, mir-499 rs3746444 and mir-149 rs2292832 and overall cancer. Further studies with a large sample size are needed to evaluate their association with cancer risk.Figure S1Begg’s funnel plot of publication bias for miR-146a rs2910164. Log OR is plotted versus standard error of Log OR for each included study. Every circle dot represents a separate study for the indicated association (C versus G).(TIF)Click here for additional data file.Figure S2Begg’s funnel plot of publication bias for mir-196a2 rs11614913. Log OR is plotted versus standard error of Log OR for each included study. Every circle dot represents a separate study for the indicated association (TT versus CC).(TIF)Click here for additional data file.Figure S3Begg’s funnel plot of publication bias for mir-499 rs3746444. Log OR is plotted versus standard error of Log OR for each included study. Every circle dot represents a separate study for the indicated association (CC versus TT).(TIF)Click here for additional data file.Table S1Scale for methodological quality assessment.(DOC)Click here for additional data file.Table S2Genotype frequency distributions of miRNA SNPs studied in included studies.(DOC)Click here for additional data file.Table S3Quality scores for included studies.(DOC)Click here for additional data file.Table S4Meta-analysis of studied miRNA polymorphisms after removing low scoring studies (score ≤5).(DOC)Click here for additional data file.Table S5Sensitivity analysis result for studied miRNA polymorphisms.(DOC)Click here for additional data file.Table S6Initial free energy (dG) predicted by mfold for the SNPs associated with the precursor and mature forms of human miRNAs.(DOC)Click here for additional data file.


CTLA-4) gene and Graves’ disease (GD) have been extensively investigated in Chinese population. However, the results were inconsistent. The objective of this study is to investigate the associations between the polymorphisms in CTLA-4 gene and the risk of GD by meta-analysis.The associations between the polymorphisms in Cytotoxic T lymphocyte-associated molecule-4  database, CNKI database and Wanfang database, covering all studies until August 11, 2012. Statistical analysis was performed by using the Revman4.2 software and the Stata10.0 software.A total of 28 case–control studies concerning the most widely studied three polymorphisms  for Chinese population in 21 publications were included. The results suggested that the G allele carriers (GG+GA) might have an increased risk of GD when compared with the AA homozygote carriers for the +49A/G polymorphism . However, as to the -318C/T polymorphism and CT60 polymorphism, the results indicated that the variant allele carriers might have decreased risks of GD when compared with the homozygote carriers .CLTA-4 gene might be risk factors for GD in the Chinese population. In future, more large-scale case–control studies are needed to validate these results.The current meta-analysis indicated that the polymorphisms in the  ADRB2ADRB2[TSHHR[CTLA-4 and IL-1-13 gene. And amoCTLA-4) is a T cell surface molecule[CTLA-4 gene is localized on chromosome 2q33. Many polymorphisms have been identified in the CTLA-4 gene. It is reported that the polymorphisms in CTLA-4 gene might influence the expression of the protein, and might play important roles in the pathogenesis of GD[CTLA-4 gene and the risk of GD. Among them, the +49A/G, -318C/T and CT60 polymorphisms were the most widely studied. To this day, the associations between polymorphisms of the CTLA-4 gene and the risk of GD have been widely investigated in the Chinese population. However, the results were inconsistent, and the associations were not yet formally evaluated. In order to derive a more precise conclusion, we performed a meta-analysis to assess the associations between the polymorphisms in the CTLA-4 gene and the risk of GD in the Chinese population. To our knowledge, this is the first comprehensive genetic meta-analysis performed in the Chinese population for Graves’ disease.Cytotoxic T lymphocyte-associated molecule-4  studies evaluating the association between the  polymorphisms in the CTLA-4 gene and Graves’ disease risk in the Chinese population, (b) the design should be a case–control design, (c) sufficient data (genotype distributions of cases and controls) available to calculate an odds ratio (OR) with its 95%CI , (d) genotype distributions in control group should be consistent with Hardy-Weinberg equilibrium (HWE). Studies were excluded if one of the following existed: (a) the studied populations were based on family or sibling pairs, (b) genotype frequencies or numbers were not presented in the original studies, (c) reviews and abstracts. If more than one study was published by the same authors using the same case series or overlapping case series, studies with the largest size of samples were included.A literature search in Pubmed database, Medline (Ovid) database, CNKI database and Wanfang database was carried out to identify studies investigating the association between the Graves’ disease risk and the Two investigators independently extracted the data and reached a consensus on all items. The following items were extracted from each study if available: first author’s name, publication year, province of origin, age of cases, genotype number in cases and controls and genotyping method.CTLA-4 gene and Graves’ disease risk was assessed by odds ratios (OR) with the corresponding 95% confidence intervals (CI). The genetic models evaluated for the pooled OR of the polymorphisms were dominant models . OR was analyzed by a fixed-effects model (the Mantel-Haenszel method) or a random-effects model (the DerSimonian and Laird method) according to the heterogeneity. Heterogeneity was assessed by a X2 based Q statistic and was considered statistically significant at p-value <0.10. When the P value was more than 0.10, the pooled OR was calculated by the fixed-effects model, otherwise, a random-effects model was used. The significance of the pooled OR was determined by the Z-test and p-value less than 0.05 was considered as statistically significant. Sensitivity analysis was conducted by sequential excluding a single study each time in an attempt to identify the potential influence of the individual data set to the pooled ORs. In addition, the possible publication bias was investigated with the Begg’s funnel plot. Funnel plot asymmetry was assessed by Egger’s linear regression test[X2 test. Statistical analysis was performed using Revman4.2 software and Stata10.0 software.The strength of associations between the polymorphisms in the CTLA-4 polymorphisms and Graves’ disease risk, abstracts, reviews or duplications of search results. After reading full-texts of the remaining 127 studies, 68 studies were excluded for not relevant to the GD risk in the Chinese population, and 9 studies were excluded for not relevant to the investigated polymorphisms . Thus, 50 studies were left for data extraction. And then, a total of 54 case–control studies were extracted for these three polymorphisms. Among 54 case–control studies, genotype numbers for control group in 7 studies were not consistent with HWE, data in 19 studies were overlapped. So these 27 case–control studies were excluded. Finally, a total of 28 case–control studies in 21 publications were identified for meta-analysis[The selection process of studies was as follows. Briefly, a total of 429 results were identified after an initial search from the Pubmed, Medline (Ovid), CNKI and Wanfang databases. After reading the titles and abstracts, 302 results were excluded for being irrelevant to analysis-27. Summanalysis-25,27,28analysis,26,28 ananalysis,16,18. TX2 was 47.22 with 16 degrees of freedom and p-value < 0.00001 in a random-effects model. Additionally, I-square value is another index of the test of heterogeneity. In Figure Z value was 5.83 . The results suggested that the G allele carriers might have an increased risk of Graves’ disease compared with those individuals with the AA homozygote. Statistically similar results were obtained after sequential excluding each case–control study for the GG+GA vs. AA comparative, suggesting the stability of our meta-analysis. Significant publication bias was detected in the funnel plot (figure not shown), and in the Egger’s test, the result was: t = 2.82, p-value = 0.013, which also indicated considerable publication bias. Summary of the results of other genetic comparisons are listed in Table A total of 4009 cases and 3651 controls from 17 case–control studies were included for data synthesis. As is shown in Figure X2 was 2.56 with 6 degrees of freedom and p-value = −0.86 in a fixed-effects model. Additionally, I-square value is another index of the test of heterogeneity. In Figure Z value was 2.18 . The results suggested that the T allele carriers might have a decreased risk of Graves’ disease compared with those individuals with the CC homozygote. Statistically similar results were obtained after sequential excluding each case–control study for the TT+TC vs. CC comparative, suggesting the stability of our meta-analysis. No publication bias was detected with either the funnel plot (figure not shown) or Egger’s test . Summary of the results of other genetic comparisons are listed in Table A total of 999 cases and 702 controls from 7 case–control studies were included for data synthesis. As is shown in Figure X2 was 0.91 with 3 degrees of freedom and p-value = 0.82 in a fixed-effects model. Additionally, I-square value is another index of the test of heterogeneity. In Figure Z value was 4.34 . The results suggested that the A allele carriers might have a decreased risk of Graves’ disease compared with those individuals with the GG homozygote. Statistically similar results were obtained after sequential excluding each case–control study for the AA+AG vs. GG comparative, suggesting the stability of our meta-analysis. No publication bias was detected with either the funnel plot (figure not shown) or Egger’s test . Summary of the results of other genetic comparisons are listed in Table A total of 818 cases and 1159 controls from 4 case–control studies were included for data synthesis. As is shown in Figure cytotoxic T lymphocyte associated-4 (CTLA-4) gene is one of the widely investigated. CTLA-4 gene, which encodes a vital negative regulatory molecule of the immune system[Graves' disease (GD) is a thyroid-specific autoimmune disease affecting 0.25–1.09% of the Chinese population. Howevere system, has beee system,32. To dCTLA-4 gene. We found that the +49A/G polymorphism was associated with an increased risk of GD in the Chinese population, and the G allele carriers might have a higher risk of disease than the AA homozygote carriers. The results suggested a significant association between this polymorphism in the Chinese population, which is consistent with some other populations, such as the UK population[This meta-analysis, including a total of 28 case–control studies in 21 publications, investigated three most widely studied polymorphisms in the pulation and the pulation. Our respulation. Thus, oA total of 999 cases and 702 controls from 7 case–control studies were included for the -318C/T polymorphism. The results suggested that the T allele carriers might be associated with a decreased risk of GD compared with CC homozygote carriers. As for the CT60 polymorphism, 818 cases and 1159 controls from 4 case–control studies were included, and the results also indicated a decrease in the risk of GD. Considering the included case–control studies for both polymorphisms were relatively small, larger number of relevant studies are needed in future to validate these results.interleukin 3 (IL3) gene was associated with increased risk of GD[interleukin-12 (IL-12) gene was also associated with increased risk of GD[ADRB2 gene[interleukin-10 (IL-10) gene[TNF-α gene[Hitherto, many studies have already been published focusing on the genetic risk factors of the GD among the Chinese population. For instance, Chu reported that a non-synonymous single-nucleotide polymorphism rs40401 (P27S) of the sk of GD; Guo fousk of GD. In addi10) gene, TNF-α gF-α gene were alsIn the present meta-analysis, sensitivity analysis was performed and stability of the results was guaranteed. Publication bias was assessed by Begg’s funnel plot and Egger’s test. No signWe have to mention the heterogeneity. We found significant heterogeneity for the +49A/G polymorphism. Since all participants were Chinese, the genetic background might not be taken as a factor for the heterogeneity for +49A/G polymorphism. However, some other factors, such as gender, age and location might affect the heterogeneity. In addition, we found no heterogeneity for the -318C/T and the CT60 polymorphisms, which suggested that the association for these two polymorphisms are more reliable than the +49A/G polymorphism.It is reported that GD occurs more frequently but less severe in women than in men. In China, the different condition of disease in men and women might be similar to the situation of the world. In our study, the data was not analyzed by gender because of the lack of original information for these populations. In future, such subgroup studies are also needed to be carried out. Moreover, the cases and controls in this meta-analysis were mostly based on Han nationality, but not in the minorities. In order to get comprehensive results of the Chinese population, studies based on the minorities are also needed.There are several limitations in this meta-analysis. First, the quantity of enrolled published studies was not very ideal, especially for the -318C/T and CT60 polymorphism. This might cause some potential publication bias, although the results of the above mentioned bias tests was not significant for these two polymorphisms. Second, data were not stratified into subgroups according to some other factors such as age, gender, location and ethnicity (Han or others), due to the lack of information in the original studies. Third, the interactions between genetic factors and environmental factors were not discussed for these three polymorphisms. Fourth, the current meta-analysis only investigated the three most widely studied polymorphisms, and some other polymorphisms with fewer reports were not included. And in future, if there were more case–control studies, new meta-analysis should be conducted. Despite of these limitations, we have minimized the bias through the whole process based on means in study identification, data selection and statistical analysis as well as in the control of publication bias and sensitivity, and got a more reliable result.CTLA-4 gene. We found that three polymorphisms  in the CTLA-4 gene were associated with the risk of GD. Our results supported the classic view that GD is associated with heredity and revealed that genes in the pathogenesis are important for GD. These results may have implications for further medicine researches about GD for the Chinese population. In future, more large-scale case–control studies are needed to validate our results.To our knowledge, this is the first comprehensive genetic meta-analysis performed in Chinese population for Graves’ disease and The authors declare that they have no competing interests.LD designed the research. JH and JQY searched the publications, extracted the data and wrote the article. YGZ checked all data. JCH and ZPX was responsible for data synthesis and helped designed the study’s analytic strategy. YGZ and LD edited the manuscript. YXM, TYX and HCW revised the manuscript. All authors read and approved the final manuscript.The pre-publication history for this paper can be accessed here:http://www.biomedcentral.com/1471-2350/14/46/prepub


Genetic polymorphisms of the Optic atrophy 1 gene have been implicated in altering the risk of primary open angle glaucoma (POAG), especially the susceptibility to normal tension glaucoma (NTG), but the results remain controversial.Multiple electronic databases  were searched independently by two investigators. A meta-analysis was performed on the association between Optic atrophy 1 polymorphisms (rs 166850 and rs 10451941) and normal tension glaucoma (NTG)/high tension glaucoma (HTG). Summary odds ratios (ORs) and 95% confidence intervals (CI) were estimated.p = 0.002; CC vs. TT: OR = 1.52, 95% CI 1.04–2.20, p = 0.029; CC vs. CT+TT: OR = 1.64, 95% CI 1.16–2.33, p = 0.005; CC+CT vs. TT: OR = 1.21, 95% CI 1.02–1.44, p = 0.032). However, no evidence of associations was detected between the OPA1 IVS8+32C>T polymorphism and POAG susceptibility to HTG. Similarly, clear associations between the rs 166850 variant and NTG were observed in allelic and dominant models  but not to HTG. In subgroup analyses by ethnicity, we detected an association between both OPA1 polymorphisms and risk for NTG in Caucasians but not in Asians. By contrast, no significant findings were noted between OPA1 variants for HTG, either in Caucasians or in Asians.Seven studies of 713 cases and 964 controls for NTG and five studies of 1200 cases and 971 controls for HTG on IVS8+4C>T (rs 166850) and IVS8+32T>C (rs10451941) were identified. There were significant associations between the OPA1 rs10451941polymorphism and NTG susceptibility for all genetic models(C vs. T OR = 1.26, 95% CI 1.09–1.47, Both the IVS8+4C>T and IVS8+32T>C variants may affect individual susceptibility to NTG. Moreover, stratified analyses for NTG detecting the effects of both OPA1 polymorphisms seemed to vary with ethnicity. Further investigations are needed to validate the association. Glaucoma, the leading cause of irreversible blindness worldwide The optic atrophy 1 (OPA1) gene In recent years, OPA1 polymorphisms have attracted widespread attention. Although several OPA1 polymorphisms have been investigated as risk factors for POAG, two polymorphisms, IVS8+4C>T and IVS8+32T>C, within the OPA1 gene have been the most extensively investigated so far. Aung et al. To date, no meta-analysis has been conducted to validate the association of polymorphisms of OPA1 with normal tension glaucoma (NTG) and high tension glaucoma (HTG). Hence, we performed a meta-analysis of all eligible studies to derive a more precise estimation of the association, to help us better understand its possible influence on POAG.MEDLINE, EMBASE, Science Citation Index, the Cochrane Library and the Chinese National Knowledge Infrastructure (CNKI) were searched up to the end of February 2012, using the key terms “OPA1 gene” or “optic atrophy 1 ” or “OPA1”, “polymorphism” or “SNP” or “single nucleotide polymorphism” or “variation” or “mutation”, and “glaucoma” or “primary open-angle glaucoma” or “POAG” or “high tension glaucoma” or “normal tension glaucoma” or “NTG” or “HTG”. Relevant publications were examined for references until no further studies were found.The following inclusion criteria were used to select literature for the meta-analysis: (1) case-control, nested case-control, or cohort studies; (2) description of the association of OPA1 polymorphisms with POAG; (3) the numbers of cases and controls and the number of different genotypes of IVS8+4C>T, IVS8+32T>C in the cases and controls were provided, or information that could help infer the results in the papers.Two observers  independently abstracted data from all eligible publications onto paper data collection forms. Two reviewers were blinded to the details  of these studies during assessment. Disagreements were resolved by discussion or consensus involving a third reviewer (HTZ) when required. The following items were collected from each study: first author’s surname, year of publication, statistical data, ethnicity, total number of cases and controls as well as numbers of cases and controls for each OPA1 genotype.The Hardy–Weinberg equilibrium (HWE) in controls was recalculated in our meta-analysis. The chi-squared goodness of fit was used to test deviation from HWE (significant at the 0.05 level). The effect measure of choice was pooled OR with its corresponding 95% CI. The assumption of heterogeneity was checked with the Q-test. A P-value less than 0.10 for the Q test indicated a lack of heterogeneity among the studies. Based on the Q-test value, two models of meta-analysis were applied for dichotomous outcomes: A fixed-effects model, using the Mantel-Haenszel (M-H method), was used to calculate the pooled ORs when the Q-test value ≥0.1. By contrast, a random-effects  model was utilized if the Q-test value < 0.1. First we compared allele frequencies  between cases and controls. We then examined OPA1 genotypes using additive , recessive  and dominant  genetic models for allele T and allele C. Furthermore, subgroup analyses were performed by ethnicity.One-way sensitivity analyses were performed by iteratively removing one study at a time to assess the stability of the meta-analysis results. Cumulative meta-analysis was performed to evaluate the accumulation of evidence on the association between OPA1 polymorphisms and NTG/HTG. Finally, publication bias was qualitatively assessed by performing Begg’s funnel plots, and it was quantitatively evaluated by Egger’s test. P<0.05 was considered representative of statistically significant publication bias. All statistical analyses were performed using a commercial statistical softwore package. Two-sided p-values <0.05 were considered statistically significant.The initial search yielded 194 articles. Based on the title, the content of the abstract and key words, 185 studies were excluded. Nine articles were reviewed in their entirety. One arm had to be excluded because of no data available The overall analyses suggested significant associations between the IVS8+32T>C polymorphism and normal tension glaucoma (NTG) susceptibility in all genetic models, and clear evidence of associations was found between the IVS8+4C>T variant and risk of NTG in allelic or dominant models . However, no evidence of associations was detected between the two OPA1 polymorphisms and high tension glaucoma (HTG) susceptibility. Because a Q-test of heterogeneity among studies was nonsignificant in all genetic models, a fixed-effects model was used.  in Asians. In contrast to the results for NTG risk in Asians, the analyses showed significant associations between both OPA1 polymorphisms and susceptibility to NTG in Caucasians . Cumulative meta-analysis of both OPA1 polymorphisms and NTG risk revealed that the summary ORs was larger than 1, and the 95% CI was reduced with accumulated sample size. By contrast, the pooled ORs of cumulative meta-analysis for HTG remained centered on 1 with increasing sample size, indicating that rs 166850 and rs 10451941 were unlikely risk variants for HTG, as shown in Publication bars were qualitatively assessed by Begg’s funnel plot and quantitatively assessed by Egger’s test. Neither Begg’s funnel plot nor Egger’s test detected obvious evidence of publication bias in the overall and subgroup analyses for all genetic models .Primary open-angle glaucoma is considered to be a complex trait with a strong genetic component In this meta-analysis, we found that the effect of the OPA1 polymorphisms on POAG risk may vary by subtype of POAG As mentioned above, the results of many studies have represented that ethnic differences may affect genetic predisposition to NTG and HTG If the effect observed in this meta-analysis was truly without bias, several possibilities may explain why such an effect was detected in Caucasians but not in Asians. On the one hand, the possible role of ethnic differences in genetic background is suggested. On the other, the relevant environmental exposure in Asia may differ from that in Western countries. Furthermore, the genetic pathogenesis of NTG may be different between the two ethnic groups. Further studies involving different racial groups from the same geographic areas may help answer this question.Some limitations in our study should be addressed, and the results should be interpreted with caution. First, controls were not uniformly defined. This study is a meta-analysis of case-control studies, most of which were hospital-based. Thus, some inevitable selection bias might exist in the results, and they may not be representative of the general population. Second, only published studies were included in this meta-analysis. Unpublished data, ongoing studies and articles published in languages other than English and Chinese were not sought, especially those with negative findings, which may have biased our results, although no obvious publication bias was apparent. Third, this meta-analysis was limited by the number of cases and controls as well as small sample sizes, especially in the subgroup analysis by ethnicity. Only a single study was conducted in Barbadian, Ghanaian, and African-American populations. Thus, additional studies are needed to evaluate the effect of these functional polymorphisms on POAG in different races. Fourth, our results were based on unadjusted estimates; a more precise analysis of the various groups should be conducted according to other factors, such as age and sex. Fifth, the genotyping methods used were different among these studies, which might have affected the results. This discrepancy between genotyping methods highlights the need for implementing rigorous quality control procedures in future studies. Finally, the existing literature lacks information about potential gene-gene or gene-environment interactions. Given that the role of several environmental factors in the pathogenesis of open angle glaucoma is established, further research should be performed in this direction.In conclusion, the results of this meta-analysis suggest that two OPA1 polymorphisms [IVS8+4C>T (rs166850) and IVS8+32T>C (rs10451941)] are associated with an increased risk for normal tension glaucoma. It is also worthwhile to note that OPA1 polymorphisms may be involved in the pathogenesis of NTG in Caucasians but not in Asians. Due to the limitations discussed above, well-designed large-scale cohorts or case-control studies are warranted to confirm our findings.


In the subgroup analysis by ethnicity (non-Asian and Asian), no significant association was observed between eNOS G894T polymorphism and MI risk among non-Asians (P>0.05), but a positive significant association was found among Asians (P<0.05).A total of 34 studies including 8229 cases and 12839 controls were identified for the meta-analysis. The eNOS G894T polymorphism is associated with increased MI risk in Asians. The results indicate that ethnicity plays important roles in the association between eNOS G894T polymorphism and MI.The  Myocardial infarction (MI) is a complex syndrome determined by multiple predisposing genetic and environmental factors. Previous studies have investigated the association of genetic variants in DNA repair pathways, lipid-related pathways, fibrinolytic system, renin angiotensin aldosterone system and nitric oxide synthase with MI risk eNOS), inducible nitric oxide synthase (iNOS). The vascular nitric oxide (NO), mainly produced by eNOS, is a critical molecule in regulating the vascular system, including the inhibition of the platelet aggregation and adhesion and reduction of vascular smooth muscle cell proliferation eNOS G894T polymorphism, a coding region variant, results in a Glu298Asp substitution and decreases the NO levels There are several forms of nitric oxide synthase such as neuronal nitric oxide synthase (nNOS), endothelial nitric oxide synthase , and Chinese Biomedical Literature Database (CBM) using the following search terms:  and  and (polymorphism or mutation or variant), without restriction on language. The included articles were published before September 2013. All eligible studies were retrieved, and their references were examined manually for other potentially relevant studies.eNOS G894T polymorphism with MI should be evaluated. c) the genotype data was available in the cases and controls. d) the control subjects must be in agreement with the Hardy-Weinberg equilibrium (HWE).The inclusion criteria were as follows: a) case-control design. b) the association of All data were independently collected from the included studies according to a standardized protocol by two investigators. The discrepancies during data extraction were resolved by consensus. The same data in different studies were used only once. The following information was extracted: first author’s name, publication year, original country, ethnicity, sample size, and number of genotype in cases and controls.eNOS G894T polymorphism and MI was assessed using crude odds ratio (OR) with 95% confidence interval (CI). The pooled ORs were determined for homozygous model (TT versus GG), heterozygous model (GT versus GG), recessive model (TT versus GT/GG), dominant model (GT/TT versus GG). The Z test was used to assess the pooled OR with the significance set at P<0.05. HWE was assessed using the Chi-square test in control groups. The presence of between-study heterogeneity was evaluated by using the I2 statistic test, which does not inherently depend on the number of studies in the meta-analysis and is preferable to the test of heterogeneity 2 ranged from 0–100%. If obvious heterogeneity was observed among the studies (I2>50%), the random-effects model (the DerSimonian and Laird method) was used to calculate the pooled OR and 95% CI The association between eNOS G894T polymorphism and MI risk were obtained in the final meta-analysis. Data collected from the included studies were summarized in the Our meta-analysis was performed according to guidelines of the “Preferred Reporting Items for Systematic reviews and Meta-Analyses” (PRISMA) statement  [14]. A eNOS G894T polymorphism and MI was found under a homozygous genetic model , a heterozygous genetic model , a recessive genetic model , a dominant genetic model  (A significant association between = 0.009) .eNOS G894T polymorphism and MI in the Asian subgroup under a homozygous genetic model , a heterozygous genetic model , a recessive genetic model , and a dominant genetic model . In contrast, no significant association was observed in the non-Asian subgroup under any of the genetic models (P>0.05)   and S4.  studies  and 4.P = 0.007, P = 0.004 respectively), which was consistent with subgroup analyses results in homozygous and recessive genetic models (P = 0.058), which was also consistent with subgroup analyses results in the dominant genetic model .P = 0.075 for homozygous genetic model; P = 0.299 for heterozygous genetic model; P = 0.118 for dominant genetic model; P = 0.055 for recessive genetic model).Publication bias of the selected articles was assessed by the Begg’s funnel plot and Egger’s test. The shape of the funnel plot did not show obvious publication bias . SimilareNOS G894T polymorphism and MI: OR = 1.41 for the homozygous genetic model, OR = 1.12 for the heterozygous genetic model, OR = 1.35 for the recessive genetic model, and OR = 1.18 for the dominant genetic model. Further stratified analysis revealed that the eNOS G894T polymorphism was significantly associated with MI in the Asian subgroup (P<0.05), but not in the non-Asian subgroup (P>0.05). The results indicated that ethnicity played important roles in the association of eNOS G894T polymorphism with MI risk.In the current meta-analysis with 8229 cases and 12839 controls, we found that there were significant associations between eNOS G894T polymorphism and the risk of coronary artery disease (CAD), MI, coronary spasms and hypertension eNOS 894TT carriers was greater than the other genotypes carriers after the exercise training eNOS 894TT genotype may have low NO in vivo and are more susceptible to endothelial dysfunction, which might increase the risk of MI. The present meta-analysis results of homozygous and recessive genetic models can account for the above hypothesis. Nevertheless, the number of TT genotype is relatively small in Asia populations and the 95%CI line of the pooled OR for Asia populations is longer than that for non-Asia population studies in A number of association studies have investigated the association between the eNOS G894T polymorphism with MI. To our knowledge, our meta-analysis represents the first one focusing on the association between eNOS G894T polymorphism and the risk of MI. In 2004, Casas et al eNOS G894T polymorphism and ischemic heart disease (IHD) including MI and CAD. They found that individuals homozygous for the eNOS 894T allele were at moderately increased risk of IHD. In 2012, the meta-analysis results of Zhang indicated that eNOS G894T polymorphism was associated with CAD risk among Asia population eNOS G894T polymorphism and MI. Our meta-analysis provided a precise result regarding the association of eNOS G894T polymorphism with MI risk.Conflicting results have been reported in investigating the association of the eNOS G894T polymorphism with MI risk. Therefore, meta-regression and subgroup analyses were performed to explore the sources of between-study heterogeneity. The results indicated that ethnicity was the source of heterogeneity in the homozygous and recessive genetic models and total sample size was the source of heterogeneity in all genetic models. Sensitivity analysis revealed that the omission of any single study did not have significant impact on the overall meta-analysis estimate. Furthermore, in the meta-analysis, funnel plot did not reflect considerable asymmetry and Egger’s test also indicated no obvious publication bias. All these made the meta-analysis results reliable to some extent.Between-study heterogeneity is common and should be explored in the meta-analysis. In the current study, significant heterogeneity was found in the association of eNOS G894T polymorphism, but the effect was not addressed in our meta-analysis.There are some limitations in this meta-analysis. First, our meta-analysis was based primarily on the unadjusted ORs with 95% CIs and the potential confounding factors were not available. Second, gene-gene and gene-environment interactions may play important roles in the function of eNOS G894T polymorphism was associated with increased risk of MI. Further stratification by ethnicity indicated the association between the polymorphism and MI was restricted in the Asians. However, large-scale studies well designed for the gene-gene and gene-environment interactions information are needed to be conducted to elucidate the associations in future.In conclusion, this meta-analysis demonstrated that the Supplement S1PRISMA 2009 Checklist.(DOC)Click here for additional data file.Supplement S2PRISMA 2009 Flow Diagram.(DOC)Click here for additional data file.Supplement S3Forest plot of myocardial infarction associated with eNOS G894T polymorphism under a heterozygous genetic model (GT vs. GG) stratified by ethnicity.(TIF)Click here for additional data file.Supplement S4Forest plot of myocardial infarction associated with eNOS G894T polymorphism under a dominant genetic model (TT/GT vs. GG) stratified by ethnicity.(TIF)Click here for additional data file.


This is the largest prospective cohort analysis to assess how dietary factors involved in one-carbon metabolism are associated with endometrial cancer incidence, using 26 years of follow-up data from the Nurses’ Health Study.The prospective cohort analysis of one-carbon metabolism dietary factors used the Cox proportional hazards model, and incorporated 788 incident endometrial cancer events from 1980 to 2006. Genotyping and unconditional logistic regression were performed on 572 endometrial cancer cases and their matched controls to examine 29 mostly non-synonymous single-nucleotide polymorphisms involved in one-carbon metabolism.MTHFR 677 TT or MTHFR 1298 CC genotype had more protective associations for many of the dietary factors and endometrial cancer, but statistical power was limited in this analysis.There were no significant dose–response relationships between intake of any of the one-carbon metabolism dietary factors and endometrial cancer incidence, but alcohol consumption of <1 drink a day was significantly protective . Those with the Dietary levels of folate, choline, methionine, vitamin B2, vitamin B6 or vitamin B12 do not appear to influence endometrial cancer incidence. Moderate alcohol intake may protect against developing endometrial cancer. Nutrients and genetic polymorphisms in the one-carbon metabolism pathway are involved in DNA methylation and synthesis, and therefore may have an important role in carcinogenesis. There is increasing evidence that abnormal DNA methylation contributes to endometrial carcinogenesis , suggestThe NHS is a prospective cohort study that began in 1976, when 121 700 female registered nurses aged 30–55 and residing in 11 US states completed an initial questionnaire. From 1989 to 1990, blood was collected from 32 826 participants.For the cohort analysis, nurses who had hysterectomy, surgical menopause or cancers other than non-melanoma skin cancer were excluded at baseline (year 1980) and each subsequent follow-up cycle. There were 1 448 170 person-years of data involving 788 cases in the cohort analysis, covering 1980 to 2006.For the nested case–control analysis, genotyping was performed on 572 cases and 572 matched controls from the NHS cohort. Matching factors included age and menopausal status. Controls were randomly selected up to and including the questionnaire cycle in which the case was diagnosed.Completion of the self-administered questionnaire and submission of a blood sample were considered to imply informed consent. The NHS protocol was approved by the Human Research Committee of the Brigham and Women’s Hospital, Boston, MA, USA.Intake level of one-carbon metabolism nutrients was calculated using validated food frequency questionnaire (FFQ) information as well as data from the US Department of Agriculture , 2004 anCases of invasive type 1 endometrioid adenocarcinoma, diagnosed between 1980 and 2006, were confirmed by medical record review. There were 788 eligible incident endometrial cancer cases during this time period who did not have missing dietary data. Among those cases, 572 had blood or buccal cell samples from which DNA was extracted for genotyping.Information on potential confounders was obtained from questionnaire responses that were updated every 2 years. Updated body mass index was calculated using height reported at baseline and weight reported at each cycle. Those missing weight in one cycle had their weight carried forward from the previous cycle, whereas those missing weight for two consecutive cycles were excluded until they again reported their weight. Smoking was quantified using pack-years.The SNPs were selected because they are non-synonymous ones that are likely to affect protein functionality in one-carbon metabolism . The minThe Cox proportional hazards model was used in the cohort analysis. The hazard ratio (HR) and 95% confidence interval (CI) were reported for risk of endometrial cancer in the categorical analysis. Tests for linear trend involved ordering the quintiles of the dietary factors and treating the values as continuous. The Anderson–Gill data structure was used to efficiently handle time-varying covariates .The unconditional logistic regression model was used in the nested case–control analysis involving SNPs in a Caucasian-only study sample, and the odds ratio was reported. The additive genetic model was used, which assumes that the effect of the heterozygous genotype is intermediate between the two homozygous genotypes. The homozygous genotype of the reference allele was coded as 0.P-values were two-sided.Analyses were done using the SAS Version 9.1 software . Quintiles were created using the rank procedure. Multiplicative interaction terms involving continuous variables were created to test for effect modification using the Wald test. All In the multivariate cohort analysis, there were no significant dose–response relationships between intake of the one-carbon metabolism dietary factors and endometrial cancer incidence . Women wMTHFR 677 and MTHFR 1298 SNPs did not significantly modify the associations between total intake of the one-carbon metabolism dietary factors and endometrial cancer incidence  and rs2886059 (P-interaction=0.03). Of the SNPs in this study, only rs202676 was significantly associated with endometrial cancer  and nutrients . It will also be informative to conduct a similar study in a population with higher intake of choline as well as plasma measurements of the nutrients.Future studies involving other populations are necessary to validate our findings, especially for the lesser studied SNPs (e.g., the non-


Genome-wide association studies (GWAS) have identified new candidate genes for the occurrence of acute coronary syndrome (ACS), but possible effects of such genes on survival following ACS have yet to be investigated.We examined 95 polymorphisms in 69 distinct gene regions identified in a GWAS for premature myocardial infarction for their association with post-ACS mortality among 811 whites recruited from university-affiliated hospitals in Kansas City, Missouri. We then sought replication of a positive genetic association in a large, racially diverse cohort of myocardial infarction patients (N = 2284) using Kaplan-Meier survival analyses and Cox regression to adjust for relevant covariates. Finally, we investigated the apparent association further in 6086 additional coronary artery disease patients.MTHFD1L was statistically significant, with a 2.6-fold mortality hazard (P = 0.007). The recessive A/A genotype was of borderline significance in an age- and race-adjusted analysis of the entire combined cohort , but this finding was not confirmed in independent cohorts (N = 6086).After Cox adjustment for other ACS risk factors, of 95 SNPs tested in 811 whites only the association with the rs6922269 in We found no support for the hypothesis that the GWAS-identified variants in this study substantially alter the probability of post-ACS survival. Large-scale, collaborative, genome-wide studies may be required in order to detect genetic variants that are robustly associated with survival in patients with coronary artery disease. Genome-wide association studies (GWAS) have identified robust genetic associations in a variety of common diseases , includiIn contrast to the identification of risk markers for incident disease, risk-stratifying patients with established disease, including those who have suffered an MI, is a cornerstone of modern cardiovascular care. Yet association studies of GWAS-generated candidate genes with prognosis following acute coronary syndromes (ACS), including unstable angina, non-ST elevation myocardial infarction (NSTEMI), and ST-elevation myocardial infarction (STEMI), require resource-intensive longitudinal designs and have rarely been performed. Although the post-ACS period is considered high risk, there is substantial heterogeneity in patient outcomes, and genetic markers could potentially be useful in defining risk, guiding treatment, and understanding the mechanisms of recurrent events and mortality. In contrast to genetic screening for incident MI, which requires screening large populations of patients without recognized disease, prognostically important genetic variations after an ACS could accelerate translation to clinical practice by focusing upon a narrower cohort of patients at high risk. Moreover, the identification of genetic pathways leading to a poor prognosis following ACS may identify new pathways of disease progression that could become novel targets for the chemoprevention of recurrent ACS.GWAS-identified risk factors for incident MI pose an important opportunity to identify genetic markers of prognosis in an ACS population, given the clinical logic that a validated risk factor for MI occurrence may also lead to more rapid disease progression after an initial event. To test this possibility, we selected as a pool of candidate prognostic markers the 95 most statistically significant of approximately 2.5 million genetic variants tested in a GWAS of premature MI occurrence  [P < 1 × 10-5) of all ~2.5 million SNPs that were included on, or imputed from, the Affymetrix 6.0 microarray and brought forward into replication stage 3 of the Myocardial Infarction Genetics Consortium Study [We tested 95 SNPs in 63 individual genes, and an additional 6 distinct gene regions containing more than one genetic locus. The 95 candidate SNPs were ranked the most statistically significant  patients. The discovery cohort was comprised of 811 self-reported white patients of European ancestry with ACS who were identified from a consecutive series of patients presenting to two Kansas City, MO hospitals , from March 2001 through June 2003. Standard definitions were used to diagnose ACS patients with either myocardial infarction or unstable angina . IndividThe Translational Research Investigating Underlying disparities in acute Myocardial infarction Patients' Health status study (TRIUMPH) served as the replication cohort, as described in a recent publication . This stMTHFD1L gene, were provided by Cleveland Clinic GeneBank  [Additional validation cohorts for survival analysis involving one SNP, rs6922269 in the arction) ,13, Emorarction) , and thearction) . ClinicaGenotyping of the 95 SNPs was performed at the Broad Institute at MIT using the iPLEX MassARRAY platform (Sequenom) on extracted leukocyte DNA (TRIUMPH) or whole genome amplified DNA (Mid-America Heart cohort) ,17. MoreP < 0.05) from Hardy-Weinberg equilibrium (HWE). Chi-square testing was used to screen for possible HWE violations, which were further investigated for statistically significant departure from HWE expectations by Monte Carlo testing involving 10, 000 random reshufflings of alleles [Genotype distributions were examined for significant deviations  to represent the study-wide statistical significance threshold [Initially, Kaplan-Meier survival analysis was performed for each variant using SPSS 17.0 . The equality of survival curves was tested by the log-rank test pooled over all genotypic strata (one degree of freedom). If the log-rank hreshold , but in P < 0.05), for a hazard ratio of 2.5 or higher, given a frequent genotype (0.5), and 80% power to detect a hazard ratio of 3.3 or higher, given an infrequent genotype (0.1) [P < 0.05, we sought to explore the possibility that null results might be related to lack of power (type II error), by examining the characteristics of the overall P value distribution by visualizing a Q-Q plot (quantile-quantile) using SPSS 17.0 .Our sample had 93% power to detect an association, by the log rank test (pe (0.1) . Given tThe clinical characteristics of the 811 cases are described in Table There were 90 deaths in the cohort, which was followed for a mean of 39.1 months, with maximum follow-up time of 60 months. Risk factor data were missing for 2 individuals. There were marked differences in overall cardiac risk factor profiles, as expected, with deceased patients having relatively advanced age, as well as more extensive cardiovascular co-morbidity, including congestive heart failure . Eight variants violated HWE at the P < 0.05 level . Three of these associations did not retain statistical significance in a Cox proportional hazards adjusted model . However, the MTHFD1L association with mortality hazard remained statistically significant following adjustment for traditional cardiac risk factors. Of these covariates, age , diabetes , prior myocardial infarction , congestive heart failure , and unmarried status  had significant associations with mortality when modeled independent of genotype. The A/A genotype (versus G/G reference) of the mitochondrial-expressed MTHFD1L (methylenetetrahydrofolate dehydrogenase (NADP+ dependent) 1-like) gene variant rs6922269 was associated with hazard ratio of 2.6 (P = 0.007). The pairwise Kaplan-Meier analysis showed significant differences for the comparison of A/A vs G/G (P = 0.01), and A/G vs G/G (P = 0.02), but not A/A vs A/G (P = 0.35), implicating the A allele as a risk factor. Of the 12 A/A individuals who died , frequent co-morbidities were hypertension (N = 9), current or former smoking (N = 9), no current alcohol consumption (N = 9), hyperlipidemia (N = 8), family history of coronary artery disease (N = 8), and type 2 diabetes (N = 7). However, the frequencies of such co-morbidities were not significantly different from the 78 other deceased patients with any other MTHFD1L rs6922269 genotype.The genotype distributions, numbers of deaths by genotypic category, and unadjusted MTHFD1L rs6922269 variant was brought forward for additional genotyping in the TRIUMPH cohort. In order to maximize power, the entire group of patients was subjected to combined Kaplan-Meier and Cox regression analysis. Characteristics, by genotype, of the entire study group appear in Table P < 0.0001). However, after adjustment for age, sex, and race, the hazard ratio for mortality in association with the A/A genotype was 1.35, with a borderline statistically significant P value of 0.052. Although sample size was too limited to document any formal trend towards increased mortality risk by number of A alleles, a suggestive decrement in risk was observed in Cox-adjusted hazard ratios, from 1.35 (A/A), to 1.21 (A/G) to 1.0 (G/G reference), as shown in Table P = 0.015) but not in Whites (P = 0.284).In order to limit multiple comparisons, only the MTHFD1L, showing high similarity of race-specific allele frequencies across sites. Each site performed a standardized Cox proportional hazards analysis, stratified by race (if applicable) and adjusted for age and sex. The A/A genotype was not associated with mortality in any of the 3 independent cohorts. In the Emory cohort, among African-Americans with CAD, the A/A genotype hazard ratio (compared to G/G reference) was 1.126 . Among whites, the HR was 0.684 . Similar results for whites were obtained in the Cleveland Clinic cohort  as well as the LURIC cohort , which had the longest follow-up of any study, with a median follow-up of 10.1 years. The corresponding median follow-up time in the Emory cohort was 2.4 years, and for Cleveland Clinc, it was 3 years.Given the equivocal, but nominally statistically significant, results in our primary replication cohort, we collaborated with independent investigators to determine the extent to which our findings might be replicable in other cohorts of patients with CAD. Raw survival data are shown in Table MTHFD1L and mortality in an initial cohort of 811 white patients with ACS. However, in the primary replication cohort, as well as three additional replication cohorts of patients with MI and/or CAD, this association was not confirmed, with a total of over 9, 000 patients studied.We initially found a nominally statistically significant association between the A/A genotype of Our experience in this study highlights that robust replication in multiple independent studies is a critical criterion for judging the validity of genetic associations. However, assembling the clinical data required to perform such replication studies is challenging, given that it may require not only the ascertainment of many thousands of patients, but also oversampling for minorities (by race and/or sex), and then tracking outcomes for many years. Our report of survival data for multiple high-priority SNPs advances the relatively neglected field of post-ACS genetic prognosis.MTHFD1L to be associated with early-onset myocardial infarction [-8 [MTHFD1L A risk allele not among the statistically significant genome-wide associations  [Although chance is the likely explanation for the apparently statistically significant association that we initially observed between mortality and rs6922269, other epidemiological factors should be considered in judging the cause-effect relationship between a putative risk factor and an outcome. In particular, prior evidence that the A allele is a risk factor for myocardial infarction is equivocal. Two major independent GWAS have reported the A allele of farction ,22. Firscy 0.28) . The A acy 0.28) .Clearly, even findings that meet accepted genome-wide criteria for statistical significance in individual studies should be replicated widely before genetic associations are accepted as valid and robust. In addition, in light of our study, it appears that realistic sample size calculations for prognostic studies may need to posit effect sizes at least as small as those that have emerged from GWAS studies of the occurrence of CAD, which are typically substantially less than 1.4. The challenge of mustering sufficient power is further amplified in prognosis studies by the fact that only a small proportion of patients will experience mortality within several years of follow-up. Therefore, sample size requirements for prognostic studies are expected to be at least as large as those for case-control studies. While it is beyond the scope of the present study to prescribe detailed sample size recommendations for future studies of prognosis in patients with common cardiovascular diseases, we would note that the CARDIoGRAM consortium included 22, 233 patients with CAD, with an approximate 3:1 control-to-case ratio. In addition to requiring large sample sizes, considerations for oversampling clinically relevant subgroups, outcome adjudication, and genotyping scope and methods are all emerging challenges to field of cardiovascular genetics as it seeks to leverage genetic risk factors to better risk-stratify outcomes in AMI suvivors. Additional limitations to this study include the heterogeneity of the present study, with respect to rs6922269 and our inability to discern whether or not there could be an effect in particular subgroups of patients, given the overall absence of a significant association.We found no convincing support for the hypothesis that SNPs identified from GWAS studies of cardiac risk are associated with all-cause mortality following ACS, suggesting that independent GWAS studies of cohorts of many thousands of ACS patients may be required in order to identify prognostic factors in biological pathways promoting post-ACS mortality.The authors declare that they have no competing interests.TMM, HK, and JAS conceived of the study, and participated in its design and coordination and drafted the manuscript. TMM and SC performed genotyping for the primary study. JH and PJ participated in the design of the study and performed the statistical analysis. HA, SLH, YP, RSP, DJE, SPW, AAQ, MK, WM, BRW, and BOB executed all aspects of the replication studies and contributed data, statistical analysis, and critical input into the drafting of the manuscript. All authors read and approved the final manuscript.The pre-publication history for this paper can be accessed here:http://www.biomedcentral.com/1471-2350/12/127/prepubTables S1-S3. Description of additional study populations. Genotyping methods in additional study populations. Supplemental results in additional study populations. Supplemental Table 1. Supplemental Table 2. Supplemental Table 3.Click here for file


MMP-1 g.-1607dupG and periodontitis susceptibility, the results remained inconclusive. The purpose of this meta-analysis was to explore its role in the development of periodontitis.Matrix metalloproteinase-1 (MMP-1) plays an important role during the destruction of periodontal tissue. Although multiple studies had focused on the association between MMP-1 g.-1607dupG and periodontitis susceptibility were included into the final analysis with definite selection and exclusion criteria. Overall and stratified analyses based on disease type, severity, ethnicity and smoking status were performed. Odds ratio (OR) and 95% confidence interval (CI) were used to evaluate the association between MMP-1 g.-1607dupG and periodontitis susceptibility, while Q test and Egger’s test were adopted respectively to assess heterogeneity among studies and publication bias.Retrieved studies from Pubmed, Web of Science, Medline and Google Scholar Search regarding MMP-1 g.-1607dupG polymorphism in homozygote  and dominant model analysis . For subgroups by type of periodontitis, increased risk of chronic periodontitis was observed on heterozygote  and dominant model . Furthermore, similar association was also detected in severe chronic periodontitis .A total of 1580 periodontitis cases and 1386 controls in 11 case-control studies were included in the meta-analysis. The pooled results showed significant association between periodontitis susceptibility and MMP-1 g.-1607dupG polymorphism was associated with chronic periodontitis, especially the severity of the disease condition.Our meta-analysis demonstrated that  Periodontitis are of the most common oral diseases around the world with high prevalence of 10%–15% Matrix metalloproteinases (MMPs), a series of proteolytic enzymes responsible for the degradation of extracellular matrix and basement membranes in the beginning and developing courses of a wide range of diseases MMP-1 polymorphisms and the susceptibility of periodontitis. The guanine addition at the -1607 position, the substitution of guanine for adenine at position -519 as well as the adenine to thymidine mutation at position -442 of the MMP-1 gene promoter were supposed to be the functional polymorphisms associated with periodontitis. However, polymorphism at position -1607  was the most extensive studied locus. During the past few years, various studies from different ethnic groups were conducted to test its relevance with periodontitis susceptibility. Nevertheless, these results still remained inconsistent, which warranted us to perform a meta-analysis to further clarify its role in the pathogenesis of periodontitis.Owing to the important role of MMP-1 in the pathogenesis of periodontitis, a variety of molecular epidemiological studies have been conducted to explore the association between MMP-1 g.-1607dupG and periodontitis risks, databases of PubMed, Web of Science, Medline and Google Scholar Search were searched  with the key words “periodontitis”, “MMP-1”  and “polymorphism” (or “variant”). The references of all identified publications were manually searched for additional studies. In the search results, only English articles were taken in. All selected studies complied with the two main criteria: (1) independent case-control study evaluating the association between MMP-1 polymorphism and periodontitis susceptibility; (2) the number or frequency of genotype given in detail. Studies with insufficient information (e.g. neither the frequency nor the number of genotype was given) were excluded. Two investigators (Li and Pan) independently extracted the data and reached a consensus in order to minimize the bias and improve the reliability. Then information including the first author’s name, year of publication, country of origin, ethnicity, type of periodontitis, source of control, number of cases and health controls, genotyping method, Hardy-Weinberg Equilibrium among controls and the main result of each publication was picked up. Different ethnic descents were classified as Caucasian, Asian or Mixed (derived from an admixture of different ethnic groups).To systematically retrieve all the case-control studies related to the association between 2 test. The association between MMP-1 g.-1607dupG and susceptibility of periodontitis was estimated by odds ratio (OR) and 95% confidence interval (CI). In addition to the overall analysis, stratified analyses based on disease type, severity, ethnicity and smoking status were performed respectively. The 2G/2G+2G/1G versus 1G/1G and 2G/2G versus 2G/1G+1G/1G comparison were estimated to assume dominant and recessive effects of the variant 2G allele respectively. The statistical significance of pooled ORs was determined by Z test. Q test based on P and I2 value was used to assess heterogeneity among studies. I2 was a value that could describe the percentage of variation across studies. The bigger I2 value, the stronger heterogeneity is. P>0.05 for the Q-test indicated no significant heterogeneity across studies, and the fixed-effects model (the Mantel-Haenszel method) was applied; if not, the random-effects model was used (the DerSimonian and Laird method) Hardy-Weinberg equilibrium of the genotype distributions among controls were estimated by a goodness-of-fit χP<0.05 was used as an indication for the presence of potential publication bias. All analyses were done with STATA software , and the P values were all two-sided.Publication bias was evaluated with the linear regression asymmetry test by Egger et al MMP-1 g.-1607dupG polymorphism were recruited and put into the final meta-analyses 11 relevant papers with 1580 cases and 1386 controls about MMP-1-519 A/G, neither found any relationship with periodontitis MMP1 -422 A/T was only mentioned in one literature There were only 2 studies for MMP-1 -1607 2G/2G homozygote was significantly associated with an increased risk of periodontitis compared with wild-type homozygote (1G/1G) . Significant association was also found in the dominant genetic model , but neither in recessive model  nor in heterozygote comparison .In general, the Appreciable differences were identified in the etiology feature between CP and AgP As showed in MMP-1 g.-1607dupG and CP by ethnicity. Similar effects were also observed in stratified analysis by smoking status.No significant association was found in Asians or Caucasians under any genetic model when analyzing the association between MMP-1 g.-1607dupG recessive model  and homozygote comparison . Z test of subgroup analysis indicated that type of periodontitis (Pz = 0.024) and smoking condition (Pz = 0.048) may be the main source of heterogeneity by homozygote comparison.Heterogeneity among studies was observed in 2 = 30.9%, P = 0.162). In addition, the pooled ORs and 95% CI under the dominant model in overall comparison was obviously influenced . . The other two studies were reserved because they did not substantially affect the heterogeneity and the results (There are three of the 11 studies at variance with Hardy-Weinberg Equilibrium  results . Then thP = 0.183 for 2G/2G+2G/1G versus 1G/1G).Egger’s test was conducted to assess the publication bias of the included studies. No publication bias was found in any genetic models. Click here for additional data file.Diagram S1PRISMA 2009 Flow Diagram.(DOC)Click here for additional data file.


FTO); however, as the number of reports increases, particularly with respect to varying ethnicities, there is a need to determine more precisely the effect sizes in each ethnic group. In addition, some reports have claimed ethnic-specific associations with alternative SNPs, and to that end there has been a degree of confusion.The pathogenesis of obesity is reportedly related to variations in the fat mass and an obesity-associated gene  and their 95% confidence intervals (CI) were estimated using per-allele comparison. Summary ORs were estimated using a random effects model.I2 test values ranging from 38.1% to 84.5%.We identified 59 eligible case-control studies in 27 articles, investigating 41,734 obesity cases and 69,837 healthy controls. Significant associations were detected between obesity risk and the five polymorphisms: rs9939609 , rs1421085 , rs8050136 , rs17817449 , and rs1121980 . Begg's and Egger's tests provided no evidence of publication bias for the polymorphisms except rs1121980. There is evidence of higher heterogeneity, with FTO may represent a low-penetrance susceptible gene for obesity risk. Individual studies with large sample size are needed to further evaluate the associations between the polymorphisms and obesity risk in various ethnic populations.This meta-analysis suggests that  Obesity is becoming a worldwide epidemic in modern society. It is prevalent in individuals of both genders and of all ages, socio-economic strata, and ethnic groups. It is estimated that the total number of overweight adults has reached more than 1.1 billion worldwide, including 312 million obese individuals and that about 10% of children are classified as overweight or obese [FTO (fat mass and obesity-associated gene) was first discovered in a genome-wide association study (GWAS) for type 2 diabetes [FTO gene was associated with obesity (or obesity-related traits) in a GWAS and a genetic association study [FTO gene, which is located on chromosome 16q12.2 and has nine exons, is strongly conserved across various vertebrate species  and emerged 450 million years ago [FTO is mainly expressed in the hypothalamus and encodes a 2-oxoglutarate-dependent nucleic acid demethylase. It may play important roles in the management of energy homeostasis [In 2007, diabetes , and, alon study ,6. The Feostasis , nucleiceostasis .FTO gene has recently attracted much attention in obesity research. Previous genetic association-based studies have shown that SNPs in the FTO gene are associated with increased body mass index (BMI) [et al. [FTO was associated with BMI, total body weight, and hip circumference; many variants have been reported to be associated with obesity, including rs9939609, rs1421085, rs8050136, rs17817449, and rs1121980. These single nucleotide polymorphisms (SNPs) lie within a 47-kilobase linkage disequilibrium (LD) block encompassing parts of the first two introns and exon 2 of FTO. The transcriptional start site of the retinitis pigmentosa GTPase regulator-interacting protein 1-like (RPGRIP1L) gene is also near the 5' end of FTO [et al. [FTO intronic SNPs and RPGRIP1L.The ex (BMI) ,10,11, aex (BMI) , glucoseex (BMI) , triglycex (BMI) , lower Hex (BMI) , waist cex (BMI) , and weiex (BMI) . For exa [et al.  showed tet al. [FTO gene with obesity (or overweight) have been replicated in large European populations [Of the SNPs that were reported to associate with obesity, rs9939609 has been of particular interest since it was discovered by Frayling et al.  The assoulations ,6. In soulations  and Oceaulations , no assoFTO polymorphisms and obesity . There are no limits on language. Two search themes were combined using the Boolean operator "and". The first theme was ("obesity" (Mesh) OR "overweight" (Mesh) OR "body mass index" (Mesh) OR "obesity, morbid" (Mesh) OR "morbid obesity" OR "morbid obesities" OR "BMI" OR "body weight" OR "leptin levels" OR "subcutaneous fatness" OR "fat mass" OR "hip circumference" OR "waist circumference" OR "lean mass" OR "body height"). The second theme was  OR "fat mass and obesity associated" (Substance Name)). Meta-analysis articles were not excluded because several original studies were often combined in meta-analysis articles on genetic association studies.We searched PubMed, MEDLINE, Web of Science, EMBASE, and BIOSIS Preview up to October 2010 for studies concerning the association between Genetic association studies and GWASs with case-control subjects in which the case subjects were obese and the control subjects were healthy were included. At least two studies had to be available for each SNP. Studies were excluded if the control subjects violated the Hardy-Weinberg Equilibrium (HWE).P-values of the HWE were calculated from the above genotype data. Adult individuals with BMI ≥30 kg/m2 were defined as case subjects of obesity, and individuals with BMI <30 kg/m2 were considered control subjects. The criteria for childhood obesity described by Cole et al. [Two reviewers (SHP and YMZ) independently searched the databases. The search results were then evaluated by five reviewers . Disagreements were resolved by discussions among the reviewers. Information on gender, author name, country, ethnicity, year of publication, mean age of examination, mean BMI , and genotypes  were retrieved. MAF and e et al.  were adoFTO and obesity risk. Therefore, the overall effects were also calculated in an additive genetic model in this meta-analysis. For comparison, the ORs in dominant and recessive genetic models were also calculated. Recently, Zintzaras [To maximize the number of studies included, analyses were conducted by calculating the ORs under per-allele comparison. We found that an additive genetic model was always employed in most of the genetic association studies concerning the association between intzaras  reportedAll statistical tests were performed with Stata 10.0 software . The effect sizes were calculated by the DerSimonian-Laird method, which is a random effect model because a modest or higher heterogeneous status was found in both allelic comparison and additive genetic models.2 test (Q test) was used to evaluate heterogeneity between studies, and a threshold P-value of 0.1 was considered statistically significant. The I2 test was also conducted to evaluate heterogeneity between studies [P-value < 0.1 considered as significant. We used the software developed by Guo et al. [P-value of 0.05 and 0.001 considered statistically significant for candidate gene association studies and GWAS [I2 >50% and much higher when I2 >75% [The Cochran's χ studies . Begg's o et al.  to test o et al. ) the sigand GWAS ,23, resp I2 >75% , and hig I2 >75% ,26. To e I2 >75% .To identify the source of the heterogeneity between studies, we performed sensitivity analyses by including and excluding some studies. Sensitivity analyses were done sequentially for all of the SNPs and all of the studies (or by some subgroups of the studies).We sub-grouped the studies into six groups  on rs9939609, four groups  on rs8050136, and two groups  on rs1121980.http://hapmap.ncbi.nlm.nih.gov/) and HaploView software  [The LD patterns of the SNPs were investigated using the HapMap Database  .et al. [et al. [P-value of less than 0.05).According to the search strategy, 170, 142, 160, 167, and 171 articles  were retrieved from PubMed, MEDLINE, Web of Science, EMBASE, and BIOSIS Preview, respectively. After the first screening, in which the abstracts or titles were read, 753 articles were excluded, and 59 articles underwent further review. After reviewing these articles, 30 additional articles were excluded, which left a total of 27 articles for inclusion in this meta-analysis. According to the PRISMA guidelines , the floet al.  . The Begg's test (P = 0.39), and Egger's test (P = 0.17) provided no evidence of publication bias.Under per-allele comparison, the OR is not available from the study by Li I2 = 85.2%, P = 0.001)  under per-allele comparison, with the largest heterogeneity in all six ethnic groups (P = 0.03)   and evidence of heterogeneity (I2 = 38.1%). Begg's test (P = 0.76) and Egger's test (P = 0.84) provided no evidence of publication bias.A total of 11 out of 12 studies reported a significant, positive association between obesity and the rs1421085 genotype Table . Also unP = 0.50).By using meta-regression, no significant correlation was found between the mean control BMI and effect size in an allelic comparison . Begg's test (P = 1.0) and Egger's test (P = 0.87) provided no evidence of publication bias.Overall, 11 studies have investigated the association of rs8050136 with obesity. Of the 11 studies, the OR  of the study by Li pe Table . A signiP = 0.60).By using meta-regression, no significant correlation was found between the mean control BMI and effect size in an allelic comparison (I2 = 45.2%). Begg's test (P = 0.85) and Egger's test (P = 0.90) provided no evidence of publication bias.All six of the studies involving rs17817449 reported a significant, positive association between obesity and rs17817449 Table . A signiP = 0.57).By using meta-regression, no significant correlation was found between the mean control BMI and effect size in an allelic comparison (I2 = 84.5%). Begg's test (P = 0.12) provided no evidence of publication bias, whereas the funnel plot (not shown) and Egger's test (P = 0.09) showed evidence of publication bias.All three studies reported a significant, positive association between obesity and the rs1121980 SNP Table . A signiP = 0.28).By using meta-regression, no significant correlation was found between the mean control BMI and effect size in an allelic comparison  and obesity risk in 41,734 cases and 69,837 controls from 59 studies, counting the cases and control subjects from each study only once. We found significant evidence for a modest increase in the risk of obesity associated with the five polymorphisms in various ethnic populations , whereas Asian populations  and South Americans have lower MAFs (between 11% and 20%). When stratifying the studies on rs9939609 into subgroups, the MAFs are different in various populations. For example, the MAF in control subjects is 0.34 to 0.44 in Caucasians, 0.11 to 0.20 in Asians, 0.31 to 0.37 in Hispanics, and 0.17 in South Americans, indicating that the MAF in Asian and South American populations is less than half of that in Caucasian and Hispanic populations. The population differences in MAF and LD structure probably have arisen through evolutionary divergence [Interestingly, we found that MAF values are very similar across all five SNPs. For example, the MAFs in the study by Price vergence .I2 = 44.6%, P = 0.008 when exclusion of the three studies vs. I2 = 44.0%, P = 0.006 when inclusion of all available studies) . These results indicate that the children's studies or GWASs are not the sources of the heterogeneity. On the other hand, with the exclusion of the study by Price et al., the heterogeneity decreased rapidly  in the Caucasian subgroup, indicating that the study by Price et al. is the main source of heterogeneity in the Caucasian subgroup. By using an additive genetic model, coupled with either the inclusion or exclusion of the study by Li et al., the I2 value in the heterogeneity test declined sharply (61.7% vs. 15.6%). Similarly, removing the study by Chang et al. caused the I2 value to increase a little (61.7% vs. 67.0%), suggesting that the study by Li et al. is the main source of heterogeneity in the Asian ethnic group.For rs9939609, we conducted sensitivity analyses. After the exclusion of the three children's studies -44, therI2 decreased sharply from 38.1% to 0.0%. Similarly, the same tendency was found with rs17817449 (data not shown), suggesting that geographic stratification in an ethnic subgroup is one of the sources of the between-study heterogeneity.All of the studies on rs1421085 are from Caucasian populations, but when the five studies were excluded  and the remaining seven studies were analyzed (six from France and one from Belgium), the r2 ≥ 0.83), and in this case, rs9939609 may be selected as a TagSNP for the other four SNPs. Similarly, in three different populations, including Caucasian, East Asian and Gujarati Indians, four of the five SNPs  are also in strong LD, indicating that rs9939609 can be used as a TagSNP for the other three SNPs.We investigated the LD structure of the five SNPs in all available ethnic populations Figure . We founet al.  for rs9939609, 1.43 (1.33 to 1.53) for rs1421085, 1.41 (1.13 to 1.76) for rs8050136, 1.54 (1.41 to 1.68) for rs17819449, and 1.36 (0.96 to 1.92) for rs1121980  differential biases due to population stratification, misclassification of clinical outcome, differences in BMI, genotyping errors and over-estimation of genetic effects in the first study and (2) differences in the pattern of LD structure across populations . Using sAnalyses of statistical power constitute a crucial step in the design of genetic association studies. Whereas conventional statistical power calculations for case-control studies disregard many basic elements of analytic complexity and can seriously underestimate true sample size requirements. Many methods have been proposed to address the sample size issue in genetic association study -60. HoweMC4R [MAF [NPC1 [Inherent limitations of this meta-analysis should be pointed out, including the following: (1) Based on the statistical tests, the heterogeneities are higher in the results of this meta-analysis. There may be many potential reasons for the higher heterogeneities, including differences in genetic susceptibility across ethnic groups and measurement error of BMI. (2) Gene-gene and gene-environment interactions could not be addressed in this meta-analysis. Actually, obesity is a complex trait, and many genes are related to obesity , such asMC4R ,62,63, MC4R [MAF , and NPCAF [NPC1 . On the AF [NPC1 -66 (3) IAF [NPC1 . (4) TheThis meta-analysis suggests that FTO may represent a low-penetrance susceptible gene for obesity risk. Individual studies with large sample sizes are needed to further evaluate the associations between the polymorphisms and obesity risk in various ethnic populations.BMI: body mass index; CI: confidence interval; GWAS: genome-wide association study; HWE: Hardy-Weinberg Equilibrium; LD: linkage disequilibrium; MAF: minor allele frequency; OR: odds ratio; ORG: generalized odds ratio; SNP: single nucleotide polymorphism.The authors declare that they have no competing interests.SHP, YMZ and MDL contributed to the design of the study. SHP, XBR, and XBL carried out the screening procedure. SHP, YMZ and FYX performed the statistical analysis. SHP drafted the manuscript. All authors read and approved the final manuscript.The pre-publication history for this paper can be accessed here:http://www.biomedcentral.com/1741-7015/9/71/prepubTable S1, Figure S1, Figure S2, Figure S3, and Figure S4.Click here for file