PPARG c.1347C>T polymorphism was correlated with cancer risk. However, past results remained controversial. In this study, we performed a case-control study on the relationship of PPARG c.1347C>T polymorphism with risk of non-small cell lung cancer (NSCLC) and subsequently carried out a meta-analysis to further assess the association between PPARG c.1347C>T and overall cancer. In our case-control study, after adjusting by age, sex, body mass index (BMI), smoking and drinking, a tendency to increased NSCLC risk was noted . In the meta-analysis, we found a significant association between PPARG c.1347C>T polymorphism and overall cancer risk . In a subgroup analysis by ethnicity, evidence of significant association between PPARG c.1347C>T polymorphism and cancer risk was found among Asians and mixed populations. In a subgroup analysis by cancer type, PPARG c.1347C>T polymorphism was associated with risk of esophageal cancer and glioblastoma. In addition, in a subgroup analysis by origin of cancer cell, evidence of significant association between PPARG c.1347C>T polymorphism and cancer risk was also found among epithelial tumor. In conclusion, the findings indicate PPARG c.1347C>T polymorphism may increase the susceptibility of cancer.Recently, several studies suggested that  It is reported that about 14.1 million cancer patients and 8.2 million cancer-related deaths have occurred in 2012 worldwide .Peroxisome proliferator-activated receptor gamma (PPARG) involves three isoforms . PPARG is an important nuclear receptor which acts as a transcriptional regulator and regulates energy metabolism . In the PPARG polymorphisms with cancer risk [PPARG NM_015869.4:c.34C>G (rs1801282 C>G) and NM_138712.3: c.1347C>T (rs3856806 C>T) polymorphisms are two common single nucleotide polymorphisms (SNPs). A meta-analysis indicated the PPARG c.34C>G polymorphism was associated with the risk of cancer in Asians [PPARG c.1347C>T polymorphism with cancer risk was not found. Several meta-analyses did not identify the association between this SNP and cancer risk [PPARG c.1347C>T polymorphism with cancer susceptibility, the obtained findings remained conflicting. In addition, the association between this polymorphism and lung cancer was not studied in Asians. Therefore, in this study, we designed a case-control study and assessed the relationship between PPARG c.1347C>T polymorphism and risk of non-small cell lung cancer (NSCLC) in Eastern Chinese Han population. Meta-analysis is a useful method of promoting the effective sample size by pooling of individual data from the enrolled studies, thus strengthening the power of the study for the assessment of genetic effects [PPARG c.1347C>T polymorphism and cancer risk more precisely, we carried out a comprehensive meta-analysis.Recently, a number of studies focused on the association of cer risk –28. PPARn Asians . Howevercer risk , 31. Alt effects . To addrP < 0.001). This study was well-matched by age and gender. The SNP information of PPARG c.1347C>T is shown in Table PPARG c.1347C>T polymorphism and Hardy-Weinberg Equilibrium (HWE) in controls.The risk factors, anthropometric data as well as demographics are listed in Table PPARG c.1347 CC, CT and TT genotypes were 57.01%, 38.00% and 4.99% in 521 NSCLC patients and 61.32%, 34.50%, and 4.18% in 1,030 non-cancer controls, respectively. The genotype distribution of PPARG c.1347C>T polymorphism is listed in Table P = 0.130). When compared with the frequency of c.1347 CC genotype, there was also no difference in the frequency of c.1347 TT genotype between the NSCLC patients and the controls . When c.1347 CC genotype was used as reference, there was also no difference in the frequency of c.1347 TT/CT genotype between the NSCLC patients and the controls . In addition, When c.1347 CC/CT genotype was used as reference, we found that there was no difference in the frequency of c.1347 TT genotype between the NSCLC patients and the controls . Adjustments for age, sex, BMI, smoking and drinking, as demonstrated in Table P = 0.106; TT vs. CC: adjusted OR, 1.20; 95% CI, 0.71–2.04; P = 0.492, CT/TT vs. CC: adjusted OR, 1.21; 95% CI, 0.97–1.51; P = 0.097 and TT vs. CT/CC: adjusted OR, 1.12; 95% CI, 0.67–1.88; P = 0.671).The frequencies of PPARG c.1347C>T polymorphism and overall cancer risk. A total of 35 abstracts were retrieved from searching of EMBASE and Pubmed databases. The selecting process of literature is presented in Figure PPARG c.1347C>T genotypes in different study are listed in Tables Next, we carried out a pooled analysis to determine the potential relationship between PPARG c.1347C>T polymorphism and the increased risk of cancer . PPARs interact with retinoid X receptors and then regulate the transcription process of many genes. PPARG has been implicated in the development of various diseases involving obesity, diabetes, inflammation, atherosclerosis and cancer –47. PPARPPARG c.1347C>T polymorphism and the development of cancer. Several individual studies have reported positive signals of PPARG c.1347C>T polymorphism with cancer risk [PPARG exon 6, a C to T substitution is a synonymous polymorphism which encodes histidine either with PPARG c.1347 C or T allele. The findings of previous epidemiological studies showed a relationship of this polymorphism with metabolic diseases such as type 2 diabetes and atherosclerosis [PPARG c.1347C>T polymorphism with NSCLC risk, and an increased risk was also found in the subsequent meta-analysis. These consistent findings demonstrated that PPARG c.1347C>TC>T polymorphism might influence the development of cancer. In the future, further evaluations with detailed environmental factors are warranted to confirm these results.With the increasing studies on genetic association, it is necessary to analyze the available data to obtain robust, replicable results. Considering the fact that a common SNP may make a small-to-moderate contribution to the risk of cancer, this pooled-analysis urges the necessity of adequate sample sizes to get a precise measurement between cer risk , 38, 40;cer risk ; howeverclerosis –53. It iPPARG gene, and did not consider other susceptibility genes or polymorphisms.Additionally, some potential limitations should be further addressed when interpreting our findings. First, the design of our case-control study was hospital-based, and the selecting bias might have occurred. Second, in this meta-analysis, the included studies based on the published studies, unpublished articles might fail to be retrieved. Third, since the significant heterogeneities were found in this meta-analysis, our findings should be interpreted with cautions. Fourth, lack of the data on environmental factors , the corresponding subgroup analyses were not conducted. Finally, we only focused on c.1347C>T polymorphism in PPARG c.1347C>T polymorphism with an increased risk of cancer, even in Asians, esophageal cancer, glioblastoma and epithelial tumor subgroups. Nevertheless, for some practical reasons, we hope that more case-control studies with the detailed environmental data to further explore the molecular mechanism of PPARG c.1347C>T polymorphism with development of cancer.In conclusion, this case-control study in Eastern Chinese Han populations, along with a comprehensive meta-analysis, identify the association of Genotyping analyses were carried out on genomic DNA of 521 NSCLC patients and 1,030 unrelated controls. All participants were come from Eastern Chinese Han population. The major included criterion of NSCLC patients were: (A) living in Eastern China area; (B) NSCLC was confirmed by pathological examination; (C) without autoimmune disease. The NSCLC patients comprised unrelated subjects who had been treated in Affiliated People's Hospital of Jiangsu University and Fujian Medical University Union Hospital. The blood samples were collected from January 2014 to December 2016. Index cases were first diagnosed with NSCLC. All patients gave a written informed consent.P = 0.268). The sex distribution of NSCLC cases and controls was well-matched (P = 0.453). According to the guidelines of Chinese blood donation, each participant was examined by a questionnaire and wrote his/her informed consent. The controls were randomly collected during the years 2014–2016. The study was approved by the Ethics Committee of Jiangsu University  and Fujian Medical University .The controls included healthy blood donors collected in the same hospitals, having the same ethnic background and similar lifestyle as the NSCLC patients. The controls were biologically unrelated to the NSCLC cases and were cancer-free. The age distribution of NSCLC cases and non-cancer controls was nearly identical  to extract the genomic DNA from the whole blood.PPARG c.1347C>T polymorphism (NP_005028.4: p.His449His) was analyzed using SNPscanTM genotyping assay . The SNP assays were confirmed by re-genotyping sixty-two (4%) randomly selected samples.t-test to examine the difference of continuous variables between NSCLC patients and non-cancer controls. In addition, we used χ2 test to determine the difference of categorical variables . HWE test in controls was undertaken using an internet-based χ2 goodness-of-fit test (http://ihg.gsf.de/cgi-bin/hw/hwa1.pl). Genotype-specific ORs with their corresponding 95%CIs and P-values were calculated by SAS 9.4 software for windows . P-values were presented using two-sided χ2-test.The continuous variables  are presented as the mean ± SD. We used Student's PPARG c.1347C>T variants and cancer susceptibility, we carried out a meta-analysis. All studies focusing on the association between this polymorphism and cancer risk were collected by searching of PubMed and Embase databases . The search was performed with the terms of (Peroxisome proliferator activated receptor gamma or PPARG) and (NP_005028.4: p.His449His or His449His or H449H or C161T or C1431T or rs3856806 or c.1347C>T) and (polymorphism or variant) and (cancer or carcinoma). Additional studies were also supplemented by a hand search of the corresponding references in retrieved articles. In this study, the language of publication was restricted to English. In our analysis, eligible studies had to meet the inclusion criteria: (1) focusing on the association between PPARG c.1347C>T polymorphism and cancer risk; (2) designed as a case-control or cohort study; (3) data could be extracted from the publications (genotypes of cases and controls); (4) published in English language; (5) genotype distribution was consistent with HWE in controls. Two authors (H. Ding and H. Qiu) extracted the detailed information from the eligible publications independently. When they met the disagreement, the third reviewer (Y. Chen) was invited to discuss every item. Finally, a consensus was reached. The following characteristics were selected and collected: the first author, year, country, ethnicity, genotyping method, cancer type, sample size, the origin of cancer cell and genotype frequencies.To further determine the relationship between P < 0.05 was defined as violation of HWE. Crude ORs with their 95% CIs were used to examine the strength of relationship between PPARG c.1347C>T polymorphism and cancer susceptibility. The pooled ORs for this polymorphism were performed under four genetic models . Stratified analyses were extensively performed with respect to origin of cancer cell, ethnicity, cancer type and quality scores. The heterogeneity across the eligible studies was tested by using a χ2-based Q-test and 2I test [2I > 50% or P < 0.1, which indicated that heterogeneity was significant [P < 0.1 was regarded as a bias. Meta-regression was conducted to analyze the source of heterogeneity [P value (two-sided) less than 0.05 were considered significant. Newcastle-Ottawa Quality Assessment Scale was harnessed to determine the quality score of the enrolled studies. If scores ≥ 7 stars, the study was defined as high-quality [For each included study, we analyzed HWE in controls using goodness-of-fit test mentioned above and  I2 test . The poonificant , 56. Othnificant . Removinnificant  and P < ogeneity . In the -quality , 60.


P < 0.00001; 2I = 83%). Additionally, asthma patients without smoking also had the increased lung cancer risk. In the subgroup analysis of race and gender, Caucasians, Asians, male, and female patients with asthma showed the increased risk of lung cancer. However, asthma was not significantly associated with lung adenocarcinoma risk. In the stratified analysis by asthma definition, significant associations were found between asthma and lung cancer in self-reported subgroup, questionnaire subgroup, and register databases subgroup. However, no significant association was observed in physician-diagnosed asthma subgroup. In conclusion, this meta-analysis suggested that asthma might be significantly associated with lung cancer risk.Some studies found that there was a significant association between asthma and the risk of lung cancer. However, the results are inconclusive. Therefore, we performed a meta-analysis. We searched the electronic databases for all relevant articles. Odds ratio (OR) with 95% confidence interval (CI) were used to calculate the strength of the association between asthma and lung cancer risk. Asthma was significantly associated with the increased risk of lung cancer (OR = 1.44; 95% CI 1.31–1.59;  Lung cancer is one of the most common malignant tumors affecting millions of people around the world. It is acknowledged that smoking is the most important risk factor of lung cancer . Sun et Asthma is one of the most common diseases of childhood, with an estimated global prevalence of 10% among children aged 6–7 . It is aThe detailed literature search strategy was showed in Figure P < 0.00001; I2 = 83%). Additionally, asthma patients without smoking also had the increased lung cancer risk . In the subgroup analysis of race, both Caucasians and Asians with asthma showed the same results . In the stratified analysis by gender, both male and female patients with asthma showed the increased risk of lung cancer . However, asthma was not significantly associated with lung adenocarcinoma risk . In the stratified analysis by asthma definition, significant associations were found between asthma and lung cancer in self-reported subgroup , questionnaire subgroup , and register databases subgroup . However, no significant association was observed in physician-diagnosed asthma subgroup . The results were showed in Table As shown in Figure P < 0.00001; I2 = 80%). In addition, after excluding the studies without adjusting smoking and age, the result was not changed . The results were showed in Table In the sensitive analysis, similar data were observed after sequentially excluding each study Figure . FurtherP = 0.683).The publication bias of the included studies was assessed by the funnel plot and Egger's test. The funnel plot was symmetric Figure . Egger'sThis present meta-analysis investigating the relationship between asthma and lung cancer risk. Eighteen studies with a total of 16375202 individuals were included in this meta-analysis. Prior asthma was significantly associated with lung cancer risk. In the subgroup analysis of race, both Caucasians and Asians with asthma showed the same results. In the stratified analysis by gender, both male and female patients with asthma showed the increased risk of lung cancer. Smoke habit was a recognized risk factor for lung cancer. However, asthma patients without smoking also had the increased lung cancer risk. These results suggested that asthma might be an independent risk factor for lung cancer. Lung cancer is classified small cell lung cancer and adenocarcinoma, squamous cell carcinoma, and large cell carcinoma . Asthma Asthma is a disease of chronic airway inflammation characterized by recurrent episodes of wheezing, dyspnea, chest tightness, and cough. Inflammatory cell types, such as T and B lymphocytes, mast cells, eosinophils, basophils, neutrophils and dendritic cells, as well as structural cell types including epithelial and mesenchymal cells involved in airway inflammation . InflammThis meta-analysis was stable and reliable. First, sensitivity analyses revealed that the results were robust. Second, no significant publication bias was found in this meta-analysis. Third, the sample size of this meta-analysis was large enough. Several limitations should be noted. First, other races, such as African, were not included. Second, heterogeneity was high in this meta-analysis. Third, subgroup analysis based on study level variables could be due to the potential for ecological fallacy. Therefore, more studies are required to confirm the results of this meta-analysis.In conclusion, this meta-analysis suggested that asthma might be significantly associated with lung cancer risk.www.wanfangdata.com.cn). We also checked American Society of Clinical Oncology (ASCO) meeting abstracts to find grey literature (http://meetinglibrary.asco.org/abstracts). Last search was updated in 10 Nov, 2016. We searched the bibliographies of identified studies and narrative reviews for additional citations. No language and time restriction were imposed in this meta-analysis. The detailed search strategy is showed in the Supplementary Material.Two authors (YLQ and JL) searched Pubmed, Embase, Chinese National Knowledge Infrastructure (CNKI) and Wanfang databases (Two authors (YLQ and JL) searched the titles and abstracts obtained from the initial electronic search for potentially relevant studies for full review. Two authors (YLQ and LXZ) then assessed the full text of the retrieved studies to determine whether the study met the inclusion criteria. Studies included in this meta-analysis should meet the following criteria: (1) study design: prospective cohort study, cross-sectional study, and longitudinal study; (2) population: individuals without lung cancer; (3) exposure: asthma or wheeze; (4) comparison: individuals without asthma or wheeze; (5) outcome: relative risk (RR), hazard ratio (HR), or OR with corresponding 95% CI of lung cancer risk in overall population and in non-smokers. If serial studies of the same population from the same group were reported, the largest study was included. Reviews, meta-analyses, letters, and editorial articles were all excluded. We resolved disagreement by consensus.Two investigators (YLQ and JL) extracted the data independently. The following data were collected from each study: first author's name, year of publication, ethnicity, gender, age, asthma definition, lung cancer definition, duration of follow-up, sample size, and adjustment. The Newcastle–Ottawa Scale (NOS) was used to assess the quality of included studies.I2 statistic and Q statistic. Subgroup analysis was carried out by smoke status, ethnicity, gender, subtype of lung cancer, and definition of asthma. Relative influence of each study on the pooled estimate was assessed by omitting one study at a time for sensitivity analysis. Sensitivity analysis was also performed by excluding studies without adjustment, studies without adjustment of age and smoking. Funnel plot and Begg's test were employed to evaluate publication bias. All statistical tests were performed using the STATA 11.0 software . A P value < 0.05 was considered statistically significant, except for tests of heterogeneity where a level of 0.10 was used. All tests were two-sided.ORs with 95% CIs were used to calculate the strength of the association between asthma and lung cancer risk. Random effects model was used in this meta-analysis. Heterogeneity among studies was examined with


Methylenetetrahydrofolate Reductase (MTHFR) polymorphisms and the risk of various types of leukemias across the lifespans of children and adults by using the meta-predictive techniques. The secondary objective was to examine the interactions among epigenetic risk factors (including air pollution), MTHFR polymorphisms, and the risks of developing leukemia. We completed a comprehensive search of 6 databases to find 54 studies  for MTHFR 677, and 43 studies  for MTHFR 1298, published from 1999 to 2014. The results revealed that, in European populations; childhood populations; children from Europe, East Asia, and America; and children with acute lymphocytic leukemia (ALL), MTHFR 677 polymorphisms  are protective, while CC wildtype was leukemogenic. In addition, MTHFR 1298 polymorphisms were protective against ALL and acute myeloid leukemia in European children, and in chronic myeloid leukemia in all adults worldwide and American adults. Air pollution played a role in the increased polymorphisms of MTHFR 677 genotypes in childhood leukemia.The major objective of this study was to examine the association between  The MTHFR is a key enzyme in folate metabolism and DNA methylation [MTHFR, C677T (rs 1801133) and A1298C (rs 1801131), have been associated with reduced MTHFR enzymatic activity, which leads to hyperhomocysteinemia, aberrant folate metabolism, and DNA hypomethylation [Leukemia is one of the most common types of hematological malignancy among children  and adulhylation . MTHFR ehylation , 9. Two hylation . Abnormahylation , 7, 8.MTHFR variant alleles against leukemia in both children and adults [MTHFR polymorphisms [MTHFR polymorphisms and leukemia risk [MTHFR polymorphism in leukemogenesis of various leukemias across lifespans of children and adults.However, other studies have reported a protective advantage of carrying d adults , 10. Leud adults . Therefoorphisms , 8, 12. orphisms  and alteorphisms –16. For orphisms –23. Outdorphisms , 19. Benorphisms , 17, 24.orphisms . Despitemia risk –27. To fmia risk , 29, we MTHFR 677 and 43 studies  for MTHFR 1298 polymorphisms. These studies were conducted in five continents, included both childhood and adulthood leukemias, and specified subtypes: acute lymphoblastic leukemia (ALL), acute myeloid leukemia (AML), chronic myeloid leukemia (CML), and various combinations of mixed types , while CC wildtype was associated with increased risk of leukemia   . Subgroup analyses by leukemia types again demonstrated the protective role of MTHFR 677 TT polymorphism against ALL in American children  while MTHFR 677 CC and CT genotypes increased risk of ALL in American children . The pooled meta-analyses did not show significant roles of MTHFR 677 polymorphisms in any of the adulthood leukemias  Table . Subgrou9) Table , Table 2as Table , Table 2MTHFR 677, MTHFR 1298 CC and AC polymorphisms played protective roles against ALL and AML in European children, cases aggregated, in 4 studies , while the AA wildtype was associated with increased risk of ALL and AML, cases aggregated, in the same studies   while AA wildtype was a risk type for developing CML  (Table Similar to 6) Table . Similar6) Table . The resMTHFR 677 polymorphisms (TT plus CT together) being protective (RR < 1), risks (RR > 1), or mixed effects (RR varied around 1). The countries that had MTHFR polymorphisms as protective genotypes (RR < 1) included mostly European countries , plus Brazil and Singapore  included Canada, Taiwan, the Philippines, India, Iran and Egypt  and together (TT and CT) played a protective role against childhood leukemia in most countries except Taiwan, China, India, Turkey, and Egypt  to visualize regional distributions . The glo regions . In the nd Egypt .MTHFR polymorphisms, we performed meta-predictive analyses using both big-data analytics and conventional analyses on childhood leukemias . To further illustrate this significance, we plotted those results on nonlinear curves. With the increased air pollution rates from low (Level 2) to high (Level 4), there was a steeper increase in the percentages of MTHFR 677 TT polymorphisms in the leukemia-case groups than in control groups (MTHFR 677 TT mutation in leukemia-case groups (more red blocks) than in control groups , MTHFR 677 polymorphisms played a protective role in European countries , had mixed effects in China and Turkey, and posed a risk for childhood leukemia in Taiwan and Iran. Among countries with low levels of air pollution (level 2), MTHFR 677 polymorphisms played protective roles against childhood leukemia in Brazil but were leukemogenic in India. These findings illustrate the complexity of gene-environment interactions across regions and ethnic groups. Air pollution has direct [MTHFR polymorphisms in leukemias. In summary, research should study how genes, environment, and additional risk factors including air pollution interact with MTHFR polymorphisms; such studies across different regions and ethnicities will advance the prevention of leukemia, especially for children.Furthermore, using meta-predictive techniques, we showed a potential impact of air pollution on the link between s direct , 19 and s direct , 17, 24 We followed the guidelines for Preferred Reporting Items of Systematic Reviews and Meta-Analysis (PRISMA) , and conMTHFR polymorphisms (C677T and/or A1298C) and leukemia risk using a case-control design, 2) included the genotype frequency in both case and control groups, 3) were written in English or 4) were written in non-English but provided tables with genotype allele frequencies for both case and control groups. Articles were excluded if they 1) were written in non-English without tables listing genotype allele counts, 2) were not case-control studies, or 3) had missing counts of genotype frequency.The inclusion criteria were that the articles 1) examined the association of the Based on those criteria, we first identified 92 articles. We excluded 38 articles, including 11 meta-analyses, 7 non-case-control studies, 8 studies on other neoplasms or carcinomas, and 12 studies that lacked genotype allele counts or duplicates. The remaining 54 papers were included in the final analyses . We calculated pooled risk ratios (RR), odds ratios (OR), and 95% CI for the MTHFR subtypes, distinguishing between cases and controls, to test for the associations of genotypes with leukemia. We compared pooled standardized RRs versus ORs to check the differences between the two ratios and found RRs more conservative with less significant findings; therefore, we used RRs in this report.® pro 13 program  for meta-prediction analyses to examine how AP death was related to the MTHFR polymorphisms and leukemia risk [p values were two-tailed with a significance level at .05. We used the GIS maps to demonstrate the distributions of mutations and their associations with leukemia risks on the global maps [We used SAS JMPmia risk . In addimia risk –43. To pmia risk . All p vbal maps , 45.p < 0.05). We performed subgroup analyses on those five studies versus those that had acceptable HWE. However, the results did not yield significant differences. Therefore, all studies regardless of HWE status were included in the final meta-analysis (We used Hardy-Weinberg Equilibrium (HWE) analysis to detect the discrepancies in genotype distributions , 47. HWEanalysis . Egger's


Genetic factors play an important role in the pathogenesis of coronary heart disease (CHD). Kinesin-like protein 6 (KIF6) is a new candidate gene for CHD, since it has been identified as a potential risk factor. The aim of this study was to perform a systematic review and meta-analysis of previously published association studies between the Trp719Arg polymorphism of KIF6 and the development of CHD.Studies and abstracts investigating the relationship between the Trp719Arg polymorphism of KIF6 and subsequent risk for development of CHD were reviewed. Electronic search from Pubmed and EBSCO databases was performed between 1993 and 2014 to identify studies that fulfilled the inclusion criteria. To analyze the association we used the models: allelic, additive, dominant and recessive. Moreover, we conducted a sub-analysis by populations using the same four models.Twenty-three studies were included in the meta-analysis. The Trp719Arg polymorphism showed a significant association with CHD when the analysis comprised the population with myocardial infarction (MI) and the additive genetic model was used. Moreover, this polymorphism showed a protective association with CHD when the analysis comprised the whole population using the recessive genetic model.Our findings indicate that the Trp719Arg polymorphism of the KIF6 gene is an important risk factor for developing MI and that allele 719Arg may have a protective association to present CHD in all populations.CRD42015024602.The online version of this article (doi:10.1186/s41065-015-0004-7) contains supplementary material, which is available to authorized users. The risk for coronary heart disease (CHD) is influenced by both environmental and genetic factors and often it is initially detected from clinical manifestations such as angina, myocardial infarction or sudden death due to artery occlusion . SeveralThe meta-analysis and systematic review were performed by following the Preferred Reporting Items for Systematic Reviews and Meta-Analyses (PRISMA) criteria , 10. TheTo perform the meta-analysis, we systematically searched for available articles in multiple electronic databases. The literature search was conducted using PubMed and EBSCO databases. Relevant studies were identified using the terms: “Kinesin 6 AND polymorphisms AND cardiovascular heart disease”, “KIF6 AND polymorphisms AND cardiovascular heart disease”, “KIF6 AND polymorphisms AND CHD”, “KIF6 AND Trp719Arg AND cardiovascular heart disease” and “KIF6 AND Trp719Arg AND CHD”. These words were combined to retrieve the summaries. The search also implicated the review of the bibliography cited at the end of the various research articles.Two researchers (González-Castro and Hernández-Díaz) working independently screened each of the titles, abstracts and full texts to determine inclusion. When the researchers were in disagreement a third researcher (Tovilla-Zárate) was consulted. Studies were included if they met the following criteria: (1) to be published in peer-reviewed journals, (2) to have a case–control study design, (3) to contain independent data, (4) to be association studies in which the frequencies of three genotypes were clearly stated or could be calculated, (5) to include diagnosis of a cardiovascular disease in the patient study group, and (6) the articles had to be written in English. Studies were excluded when: (1) they were not case–control studies, (2) they were reviews, comments or editorial articles, (3) provided insufficient data, and (4) they were repeated studies.The same authors mentioned previously extracted the information from all the included reports and reached consensus on all the items. The following data were obtained from each of the studies: authors, year of publication, location, ethnic group, number of cases and/or controls, age, gender and cardiovascular diagnosis of the participants. If these data were not available in the studies, the corresponding author of the respective article was contacted.http://dxsp.sergas.es) used for this part is freely available for epidemiologic analysis of tabulated data. On the other hand, to explore the problem of publication bias, the Egger’s test and funnel plots were calculated with the same software. This last approach standardizes the effect of each of the published studies on the vertical axis and its correspondent precision on the horizontal axis. Sample heterogeneity was analyzed with the Dersimonian and Laird’s Q test. Q test results were complemented with graphs to help the visualization of those studies favoring heterogeneity. For all these procedures we used the above-mentioned program. Moreover, to aboard the problem of a small sample size we performed a meta-regression based in ages; this analysis was carried out in the comprehensive meta-analysis software version 2. Next, a chi-squared (χ2) analysis was used to calculate the Hardy-Weinberg equilibrium to evaluate genotype distribution. In order to strengthen the analysis we evaluated publication bias by using the GRADE approach and assessed the risk of bias. The Newcastle-Ottawa Assessment Scale (NOS) was used for inclusion in the systematic review by scoring the methodological quality. We established a score of six as cut-off point to distinguish high from low quality studies [For the meta-analysis, the odds ratio (OR) and 95 % confidence interval (CI) values were estimated and used to evaluate the strength of the association of KIF6 Trp719Arg polymorphisms with CHD risk. Pooled ORs were calculated following four genetic models: dominant (A/G + A/A vs G/G), recessive (G/G + A/G vs A/A), additive (A/A vs G/G) and allelic (A vs G). The EPIDAT 3.1 program ( studies  (see Addn = 23) were published between 1993 and 2014.The electronic searches yielded 34 potentially relevant studies. From these 6 reports were excluded because they consisted of duplicated publications, hence 28 studies were potentially relevant for inclusion in our study; 5 studies were further excluded because either the Trp719Arg genotype was not detected or the studies did not present a control population. In the end, a total of 23 articles met the inclusion criteria  was available in 5235 patients. The analysis of CAD populations did not show a significant association between the Trp719Arg polymorphism of the KIF6 gene and CHD using the random effects model in all four genetic models    Table  and 11.FIn this study, we explored the relationship between the Trp719Arg polymorphism of the KIF6 gene and CHD given that several lines of evidence have shown an association between the Trp719Arg polymorphism and increased risk of CHD in the placebo groups of some clinical trials such as the Cholesterol and Recurrent Events (CARE) study and the West of Scotland Coronary Prevention Study (WOSCOPS) . MoreoveOur findings demonstrate that this polymorphism may be a risk for heart disease development. In addition, some limitations of the present meta-analysis must be addressed. First, we performed the present meta-analysis based only on published studies. We consider that by selecting only published results, we ensure that our meta-analysis excludes poorly designed studies. Second, although the present analysis involves 23 studies, it is relatively small in comparison with other meta-analyses on different diseases. However this limitation involved a meta-regression analysis in order to resolve the problem. We consider that the number of eligible studies included in our meta-analysis is small, hence to validate our results a larger number of studies must be included in future investigations. In the sub-group analysis by ethnicity, we only included Caucasian populations but we acknowledge the importance of theTrp719Arg polymorphism for CHD development in Asian populations. We suggest that more studies investigating this association must be undertaken in Asian populations.The present study has demonstrated that the Trp719Arg polymorphism of the KIF6 gene is an important risk factor for developing MI. Moreover, our findings suggest that allele 719Arg may exert a protective association to present CHD in all populations.


FCGR2A rs1801274 polymorphism and the susceptibility to autoimmune diseases more precisely.The aim of this meta-analysis was to estimate the association between the FCGR2A gene variants and ADs by allelic contrast, homozygote contrast, the recessive model, and the dominant model.A meta-analysis was conducted on the association between the FCGR2A rs1801274 polymorphism were found in KD  and UC . A overall meta-analysis increased risk of AD significant association between FCGR2A rs1801274 gene polymorphism and ADs under allelic , homozygous , dominant  and recessive  in Asian population. Meanwhile, a decreased risk of AD was detected in the allelic , homozygous , dominant  and recessive  in African-American population.A total of 17 studies with 30 comparisons in different populations and genotype-methods were available for this meta-analysis, including 10 Kawasaki disease (KD), 7 Ulcerative colitis (UC), 6 Crohn's disease (CD), 3 Rheumatoid arthritis (RA), 2 Systemic lupus erythematosus (SLE), 1 Autoimmune thyroid disease (ATD) and 1 diabetes mellitus type 1 (T1D). A significant association between FCGR2A rs1801274 G-allele confers susceptibility to KD and UC. Data also suggests that the FCGR2A rs1801274 polymorphism may be associated with the susceptibility of multiple ADs in Asian and African-American populations.This meta-analysis demonstrates that the  Autoimmune diseases (ADs) are some complex disorders includes a substantial portion of pathobiology, the potential inheritance and induced environment, efficacy of the common therapies and co-occurrence among diseases. ADs accounted for 4-5% of the population and had always become difficult issues . More reFCGR2A had been reported and it was associated with the susceptibility to multiple autoimmune diseases, including systemic lupus erythematosus (SLE), Kawasaki disease (KD), diabetes mellitus type 1 (T1D), autoimmune thyroid disease (ATD), ulcerative colitis (UC), Crohn's disease (CD), rheumatoid arthritis (RA) FCGR2A rsluded KD , 27 met FCGR2A rs1801274 A-allele and KD  and UC . In KD, significant association were observed in the population of Caucasian  and Asia population. For UC, significant association was found in Asia population  , homozygous , dominant  and recessive  were found. In African-American population, decreased risk of AD was detected in the allelic , homozygous , dominant  and recessive . However, no significant association was found when all the contrasts were performed in Caucasian and non-Caucasian Europe populations.The summary of meta-analysis for the P value for the distribution of genotype for HWE in the control groups was calculated. Only one study in control population was not consistent with HWE. However, when studies were omitted individually from the meta-analysis to evaluate possible individual influences, heterogeneity disappear in CD and RA.As shown in Table FCGR2A rs1801274 A-allele and CD in Caucasian population. The study with the highest OR were then excluded, and the heterogeneity disappeared but the result remained none significant .Then we pooled results according to the ethnicity of study populations. Heterogeneity was found for the Publication bias analysis on KD, UC, CD and overall phenotypes were performed. but not on RA, SLE, ATD and T1D due to small numbers of these studies. No obvious asymmetry evidence was detected according to the shapes of the funnel plots (data not shown).No evidence of publication bias by the method of Egger's test and Begg's tests Table  were fouADs invokeed a wide spectrum of signs and symptoms, genetic analysis revealed multiple genes underlying distinct autoimmune conditions, which provided direct evidence that might have common immunopathologic mechanism.FCGR2A gene was located on chromosome 1q23, which consists of 7 exons spanning approximately 18.58 kb of genomic DNA. It encodes a member of a family of Fc γ receptors for immunoglobulin G. FCGR2A protein played an essential role in the protection against foreign antigens by removing antigen-antibody complexes in the circulation, and transduces activating signals into cells via immune receptor when ligated with immune complexes [FCGR2A and FCGR3A activate immune cell functions, including phagocytosis, and the release of inflammatory mediators, were therefore linked to the pathogenic consequences which triggered by autoantibodies or immune complexes in multiple immune-mediated diseases [omplexes . This stdiseases .FCGR2A rs1801274 was a missense variant leading to an amino-acid substitution of histidine by arginine at position 131 (H131R). The variant H131R was reported to interact differently with immunoglobulin G subclasses, and associated with the susceptibility to multiple autoimmune diseases. Specifically, the substitution of FCGR2A rs1801274 affected the Fc region of IgG receptor and determined the affinity of FCGR2A for IgG subclasses. This variant was capable of binding to and mediating phagocytosis with IgG2, potentially leading to altered immune response to infectious agents and activation of B cells and overproduction of cytokines [The associated polymorphism ytokines .However, owing to low prevalence or the inadequate statistical power, the discrepancy of the results was always achieved among different studies. Meta-analysis was a method of increasing the effective sample size through the pooling of datas from individual studies, thus enhancing the statistical power of the analysis for the estimation of genetic effect .FCGR2A rs1801274 polymorphism and autoimmune diseases including KD and UC, indicating a protecting effect to KD and predisposing to UC, FCGR2A rs1801274 polymorphism might have no effect on CD or RA. However, the exact biological mechanism that the FCGR2A gene polymorphisms influence susceptibility to ADs remains unclear.The results of the meta-analysis revealed significant association between the FCGR2A rs1801274 polymorphism susceptibility to ADs in the overall population was failed to detect. Sub-population analysis revealed association between different genotypes and Asia population. The AD risk increased significantly in Asia populations using allelic, homozygous, recessive and dominant genetic models. Further studies including larger sample size, well-designed case-control studies in different ethnic populations will be needed to unravel their exact roles in the pathogenesis of multiple ADs.The significant association between Several limitations of the present meta-analysis should be showed. First, for several autoimmune diseases, the number of studies is small, and this might cause insufficient power to detect slight association. Second, significant heterogeneity between-study was found in some comparisons. Third, the majority of the included studies were conducted in population of Caucasian and Asia, thus further studies in other ethnic populations were required. Fourth, the present meta-analysis was based on uncorrected estimates. A more precise analysis could be performed if the potential confounding factors including sex, age, environmental factors and other lifestyle factors were available. Thus, the results of the meta-analysis should be interpreted with caution.FCGR2A rs1801274 polymorphism was associated with the susceptibility to multiple autoimmune diseases including KD and UC. It also provided evidence that the FCGR2A rs1801274 polymorphism may be associated with susceptibility to multiple ADs in Asia population. The FCGR2A rs1801274 polymorphism in multiple autoimmune diseases provides further evidence supporting the concept of common gene underlying multiple autoimmune diseases. Further studies including larger sample size, well-designed case-control studies in different ethnic populations will be needed to unravel their exact roles in the pathogenesis of multiple ADs.In summary, the meta-analysis demonstrated that the FCGR2A rs1801274 polymorphism with autoimmune diseases were conducted. by searching the following Medical Subject Heading (MeSH) terms and/or text words: “Fc fragment of IgG, low-affinity IIa, receptor”, “FCGR2A”, “polymorphism”, “autoimmune diseases”, and “autoimmunity” in PubMed and Web of Science literature base for the period up to 2015-12-01, and relevant articles were identified for further literature filtering. A study was included if the following criteria were satisfied (1) case-control genetic association study published before June 2015; (2) genotype frequencies in the cases group and controls group were both available for estimating an odds ratios (OR) and their 95% confidence interval (CI). The exclusion criteria included: (1) studies that contained overlapping data with other literatures; (2) studies which the genotype or allele frequency could not be obtained; (3) data came from case-reports, reviews or abstracts; (4) control group did not confirm to Hardy-Weinberg equilibrium (HWE).A comprehensive search examining the association between the FCGR2A rs1801274 polymorphism. Data extraction was performed independently by two authors, and disagreements were solved by discussion.The following information from each study were extracted: first author's name, publication year, country, ethnicity, genotype-methods, the number of cases and controls and allele frequencies of the versus A , GG versus AA (homozygote contrast), G/G+G/A versus AA (dominant models), G/G versus A/A+G/A (recessive models). The heterogeneity between studies was evaluated by Cochrane's Q-statistic as well as I2-statistic which was used to quantify the effect of heterogeneity (I2 = 100% (Q-df)/Q) [Statistical analysis for the meta-analysis was conducted by Stata version 10.0 . This meta-analysis examined the contrast of G Q-df)/Q) . Subgrout-test suggested by Egger, P value less than 0.05 was considered significant publication bias.The funnel plot was used to assess potential publication bias . Egger's


Haemophilus influenza type B (HiB) vaccine. We observed a protective association between any vaccination in the first year of life and risk of childhood leukemia (summary odds ratio (OR) 0.58 [95% confidence interval (CI) 0.36–0.91]). When individual vaccines were analysed, some evidence of an association was seen only for BCG (summary OR 0.73 [95% CI 0.50–1.08]). In conclusion, early vaccination appears to be associated with a reduced risk of childhood leukemia. This finding may be underpinned by the association observed for BCG. Given the relatively imprecise nature of the results of this meta-analysis, our findings should be interpreted cautiously and replicated in future studies.Leukemia is the most commonly diagnosed childhood cancer, although its etiology is still largely unknown. Growing evidence supports a role for infection in the etiology of acute lymphocytic leukemia (ALL), and the involvement of the immune system suggests that vaccination may also play a role. However, the findings presented in the published literature are inconsistent. Therefore, we conducted a PRISMA systematic review and meta-analysis. 14 studies were identified and meta-analyzed. Vaccinations studied comprised Bacillus Calmette-Guérin (BCG) vaccine, Triple vaccine, Hepatitis B vaccine (HBV), Polio, Measles, Rubella, Mumps, trivalent MMR vaccine and  Acute lymphocytic leukemia (ALL) and acute myelogenous leukemia (AML) are the most common subtypes. ALL alone accounts for 26% of cancers diagnosed from birth until four years of age. On the other hand, chronic leukemia is considered to be very rare in this age group2. Despite the great success in the treatment of ALL where the survival of children has steadily improved reaching 85% in both the United States and Europe, its etiology is still largely unknown3. A recent study by Ward et al. showed that the incidence of childhood acute leukemia in the United States had an annual increase rate of 0.7 percent for ALL and 1.1 percent for AML between 1975 and 20102. However, the cause of this rising incidence has not been identified. Improved diagnostic techniques and improved notifications may partly explain these observed increases.Leukemia is the most commonly diagnosed cancer in childhood, representing 31% of all cancer diagnoses5, paternal smoking8, exposure to household pesticides9 and benzene10. On the other hand, early exposure to infections may be protective for childhood ALL. This was originally postulated by Greaves in 1997 when he stated that both the pattern and timing of infections in early life is critical to the developmental programming of the immune system11. These immunomodulatory changes start with the passage of the mother’s immunoglobulins through the placenta in early pregnancy and in breast milk which is considered to be the first exposure to bacteria and viruses. He argued that current lifestyles in developed countries characterized by less exposure to infections or early weaning prevent mothers from sufficiently providing immune modulation to their offspring. Moreover, the improvement of hygiene in developed countries decreased the exposure of children themselves to infections. Therefore, when the inappropriately programmed immune system in a genetically predisposed child is exposed to common microbes in an abnormal time frame, it causes an abnormal immunological response thus increasing the risk of leukemia11. A recent meta-analysis of 14 case-control studies using day-care attendance as an indicator of the increased likelihood of early exposure to infections has shown a strong protective association between day care attendance and risk of ALL12. The growing evidence in support of the role of infection in the etiology of childhood leukemia has led to an interest in a possible role of childhood vaccination.Various risk factors for childhood leukemia have been identified, such as ionizing radiation during prenatal and postnatal life18, Canada20 and Finland22 have reported a reduced risk of leukemia with Bacille Calmette–Guerin (BCG) vaccination. However, a study in New Zealand23 showed no effect. After a thorough review of the literature, we aimed to systematically review and meta-analyze the current literature to provide evidence regarding the association between a range of vaccines and the risk of childhood leukemia.Published studies have reported contradictory results; studies from USA24  and SIGLE (System for information on grey literature in Europe). The descriptor used for searching in eight databases, except Google Scholar was: “(immunization or vaccines or vaccine or vaccination or vaccinated or immunized) and (risk or risks or association or correlation or odds) and (leukemia or leukaemia or leukemic or leukaemic) and (child or children or childhood or baby or infant or infants or newborn or neonates or pediatrics or pediatric)”. In Google Scholar, we used the keywords as the exact phrases “childhood leukemia”/ “childhood leukaemia” and “immunization, vaccines, vaccine, vaccination, vaccinated, immunized” with the option with at least one of the words and selected “where my words occur in the title of the article”. For every database, all variants spelled with “s” instead of “z” were covered in the search strategy. In addition, we manually searched the reference lists of relevant articles.Our study was conducted according to the recommendations of the PRISMA statementStudies that reported the association of early vaccination with leukemia in children under 20 years old and vaccines were included in this review with no restriction to language. Exclusion criteria were: (i) unreliably extracted data; (ii) leukemia cases reported in adults; (iii) overlapped data sets; (iv) abstract-only articles, review articles, theses, conference proceedings, books, and case reports. All studies identified were reviewed independently for eligibility by three authors and conflicts were resolved by seeking a discussion and consensus with senior researchers .Three independent reviewers extracted the data from the included studies. Any discrepancies were resolved by discussion to reach a consensus or through consulting senior researchers. A form for data extraction was designed for the authors to use. Sample: , exposure assessment, statistical analysis and risk estimation through the reported odds ratios (OR) and relative risks (RR). Where these had not been estimated and only raw numbers of participants were provided in tables, we estimated crude ORs from the information available.26 was independently assessed by two reviewers using “The Cochrane Collaboration’s tool”. It consisted of six specific domains, including: sequence generation, allocation concealment, blinding of participants and personnel, incomplete outcome data, selective outcome reporting, and other sources of bias. The assessment of each domain was categorized as ‘low risk,’ ‘high risk,’ or ‘unclear risk’ of bias by two independent reviewers. Any differences between reviewers were resolved by discussion and by consultation from supervisor to reach a consensus if needed. Methodological quality assessment of observational studies32 was performed by using the risk of bias assessment tool of the “Cancer Council Australia”33. The metrics of this tool were as follows: subject selection , comparability of groups on demographic characteristics and important potential confounders, ascertainment of exposure/treatment, and follow-up.The risk of bias of clinical trial35. Random and fixed effects models were used depending on the heterogeneity among the studies.Quantitative synthesis with meta-analysis was performed using Comprehensive Meta-analysis software version 2.0 . Summary odds ratios (ORs) and 95% confidence intervals [CIs] were computed for individual vaccines when there were two or more studies providing results for the association between that specific vaccine and risk of childhood leukemia. Heterogeneity between studies was assessed using the Q statistic and I-squared tests. Heterogeneity was considered statistically significant if the P-value was < 0.1 or I-squared value was > 50%36. Publication bias was evaluated by funnel plot and Egger’s regression where there were five or more studies assessing the effect of a particular variable. Publication bias was considered significant when the P-value was < 0.1. The Duvall and Tweedie’s trim-and-fill method was used to increase symmetry by adjusting for studies that appeared to be missing in the presence of publication bias39.When there were ten or more studies of a particular vaccine, we performed meta regression for moderators to examine the potential effects of study characteristics on the summary effect measure. Subgrouping of early/late and type of vaccination was performed to examine the effect of age at vaccination and type of infection on risk of leukemia. The effect of study characteristics on the summary OR was considered significant when the p-value was less than 0.0541, two studies enrolled patients at more than 20 of age43, one study had overlapping participants44 and one study was a conference paper45. Hence, the total number of articles for final analysis was 14. Detailed characteristics of included studies are shown in Table 32, and two were clinical trials26 , but no publication bias was seen 32. Using a random-effects model, the summary OR was 0.73 [95% CI 0.50–1.08] 26 were associated with a lower risk of childhood leukemia, although the latter estimate lacked precision as there were only two studies. There was no evidence of association when the timing of vaccination was not reported31 28 revealed no association between HiB vaccination and childhood leukemia (summary OR 1.06 [95% CI 0.64–1.76])    vaccines. Meta-analysis of four studies29. The summary OR was 0.98 [95% CI 0.70–1.39]  . A subgroup meta-analysis was performed to investigate whether the association was affected by type of injection . No evidence of an association was seen in either subgroup (summary ORs were 0.84 [95% CI 0.39–1.83] and 1.02 [95% CI 0.66–1.59] respectively) . Subgroup analysis indicated that rubella vaccination was not associated with leukemia risk whether given alone or in combined trivalent MMR vaccine  vaccines – appeared to be associated  with a reduction in leukemia risk.The protective association we observed between early vaccination (and/or BCG) and risk of childhood leukemia overall may involve similar biological pathways to those underlying the apparent reduction in leukemia risk associated with day-care attendance54; outcomes for which there is some evidence of a protective include all-cause infant mortality53 and childhood cancer32. The biological pathways involved are unclear, but are thought to involve lymphocyte activation and/or innate immune memory that could promote protection beyond the intended target pathogen. Potential mechanisms are discussed in detail by Goodridge et al.53. In addition, many animal studies have reported beneficial effects of BCG in enhancing the immune response to unrelated pathogens55 and tumours. Theorically and practically, BCG vaccine has been utilized as an immunotherapy for a certain cancer58. The mechanism in which BCG vaccine ameliorates effectively cancer was not well-explored. Several studies investigated the alteration of different types of cytokines by BCG vaccine62, such as Th1’s cytokines  and Th2’s cytokines . These substances play roles in regulation of natural killer cells, macrophages or cytotoxic T lymphocytes cells61, which have antitumor effects. In addition, the antitumor activity of BCG vaccine through activating human dendritic cells, an antigen-present cells, has been studied67. This process was mediated by modulation of several cytokines such as IL2, IL10 and IFN gamma67. Taking specific antileukemia effect into consideration, studies were performed and their findings were promising. Jang et al.65 investigated the activity of dendritic cells of pediatric leukemia patients exposed by BCG, the result revealed an enhanced proliferation and maturation status. Recently, a BCG-derived heat shock protein, named HSP70, was investigated. Findings from animal and in vitro studies demonstrated an elevated immunogenicity effects against leukemia cells69.Heterologous (or “off-target”) effects of BCG vaccination in children have been consistently observed70 and reported according to the recommendations of the PRISMA statement24. Moreover, stratification according to time of vaccination (for BCG vaccination) and type of vaccination  was further examined.To the best of our knowledge, this is the first comprehensive systematic review and meta-analysis investigating the association between vaccination and childhood leukemia. All steps of this systematic review and meta-analysis were conducted in strict accordance with the Cochrane handbook of systematic reviews22 which was at moderate risk of bias, the six remaining studies were at high risk of bias. However, these biases (if present) would most likely have similarly affected the results for all types of vaccinations, not just early vaccination and/or BCG itself. We acknowledge that the association seen with BCG vaccination was not statistically significant; however, this was likely due to a lack of statistical power and a relevant protective association remains a possibility. While seven studies reported early vaccination, four studies were before 1990 and only two were in 21st century; this led to some heterogeneity. Most studies reporting BCG vaccines were conducted before the 1980s. The design of those studies was not appropriate for evaluating the effect of BCG vaccination in risk of leukemia, because clinical trials were predominantly designed for protecting against tuberculosis. Additionally, three of them, which contained a large number of participants, reported leukemia deaths rather than leukemia cases27.Although our systematic literature search was conducted on nine electronic libraries, the number of studies included in the final analysis was limited and it was not possible to conduct our meta-analysis by leukemia subtype. Furthermore, most of the included studies were observational. We were unable to evaluate the effects of several confounders such as prenatal and postnatal care, the nutrient supplies, and the microorganism exposure on the pooled results due to the lack of such information in the included studies. Further, some of the observational studies provided raw data only, from which we estimated crude study-specific ORs. Observational studies are subject to biases such as selection and recall bias. Except for Salonen 1976Haemophilus influenza type b. These changes may alter the potential effect of early vaccine on childhood leukemia risk. Further studies should investigate the impact of these shifts in vaccine policy.The change in vaccine schedule is worthy of note. According to Center for Disease Control and Prevention (CDC), many vaccines have currently been used for children at birth, such as HBV, or within the first year, such as DTA, Rotavirus, or Our comprehensive systematic review and meta-analysis indicated that the early vaccination may reduce the risk of childhood leukemia, and this finding may be attributable to BCG vaccination. Replication of this finding is important, and its impact in the context of changing vaccination policy should be carefully evaluated.Supplementary Information


A risk prediction model of non-syndromic cleft lip with or without cleft palate (NSCL/P) was established by a discriminant analysis to predict the individual risk of NSCL/P in pregnant women.A hospital-based case–control study was conducted with 113 cases of NSCL/P and 226 controls without NSCL/P. The cases and the controls were obtained from 52 birth defects’ surveillance hospitals in Hunan Province, China. A questionnaire was administered in person to collect the variables relevant to NSCL/P by face to face interviews. Logistic regression models were used to analyze the influencing factors of NSCL/P, and a stepwise Fisher discriminant analysis was subsequently used to construct the prediction model.P < 0.001). Self-verification showed that 83.8 % of the participants were correctly predicted to be NSCL/P cases or controls with a sensitivity of 74.3 % and a specificity of 88.5 %. The area under the receiver operating characteristic curve (AUC) was 0.846.In the univariate analysis, 13 influencing factors were related to NSCL/P, of which the following 8 influencing factors as predictors determined the discriminant prediction model: family income, maternal occupational hazards exposure, premarital medical examination, housing renovation, milk/soymilk intake in the first trimester of pregnancy, paternal occupational hazards exposure, paternal strong tea drinking, and family history of NSCL/P. The model had statistical significance (lambda = 0.772, chi-square = 86.044, df = 8, The prediction model that was established using the risk factors of NSCL/P can be useful for predicting the risk of NSCL/P. Further research is needed to improve the model, and confirm the validity and reliability of the model. Non-syndromic cleft lip with or without cleft palate (NSCL/P) is the most common craniofacial congenital anomaly. The incidence of the anomaly worldwide is 0.3 to 1.9 per thousand live births , which iDue to the complicated pathogenesis of the disease, the etiology of NSCL/P has not been fully understood, and the existing evidence today suggests a multifactorial inheritance for this anomaly, with both genetic and environmental causal factors. Recently, most studies have focused on the identification of risk factors of NSCL/P. Many epidemiological studies have confirmed that maternal age –12, matePrevious studies have shown that a statistical prediction model based on the risk factors is an effective method for predicting the individual risk of disease, such as coronary heart disease, hypertension, and type-2 diabetes mellitus (DM) –32. For However, there is rare research about individual risk prediction of birth defects. In our previous study, we used a decision tree to predict the risks of total birth defects and congenital heart disease based on risk factors in the first trimester of pregnancy , 34. TheWe conducted a hospital-based case–control study on mothers whose fetuses or neonates were between the 28th week of gestation and the 7th day after birth  and were diagnosed with non-syndromic cleft lip with or without cleft palate (NSCL/P) between July 2012 and June 2013 in 52 birth defects’ surveillance hospitals in Hunan Province, China. Mothers who delivered normal infants at the same hospitals as the cases were randomly selected as the controls. Additionally, the interval of the birth dates between the normal infants and the patients with NSCL/P was no more than 1 month. Those mothers were aged 20–45 years. The diagnosis of NSCL/P was performed by the clinical geneticists of those birth defects surveillance hospitals. Infants with chromosomal anomalies and other birth defects of known aetiology were excluded from the survey. Infants with cleft palate only were also excluded from the study. Those who could not cooperate with the survey were excluded from the study.In this hospital-based study, the control-to-case ratio was 2:1, due to the relatively small number of cases and a large number of potential controls to be selected from the birth defects’ surveillance hospitals. In case of few cases, using the control-to-case ratio of 2:1 could ensure the necessary statistical power to identify important predictors.The survey was conducted by obstetricians and gynecologists who were also trained investigators using the unified questionnaire with the participants in person by face to face interview. The unified questionnaire was designed by the experts on our research team, and was modified based on the pilot study. The contents of questionnaire were classified 5 categories and 28 variables, including sociodemographic characteristics of the mothers, economic status of their families, family histories, conditions of the mothers from 6 months before conception through the first trimester of pregnancy and characteristics and conditions of the fathers.Sociodemographic characteristics and family income Maternal age was classified into four scales (years): 20–24, 25–29, 30–34, ≥35. Maternal education level was classified into three categories: primary school and below, middle school, college and above. Maternal occupations included farmers, migrant workers, employers/managers, workers, staffs in administrative institutions, and housewives or else. Family income was classified into four scales (yuan/year/person): ≤5000, 5001–10,000, 10,001–15,000, >15,000.Family histories Family histories of NSCL/P were defined as one or more first relatives of one person suffering from NSCL/P. In this study, family histories of NSCL/P were included the family histories of mother and father. Abnormal reproductive histories referred to the histories of stillbirth, spontaneous abortion, or birth defect.Conditions of the mothers In this study, most variables were dichotomies, collected from the questionnaire using the questions with answers yes or no, including occupational hazards exposure, premarital medical examination, chronic disease, upper respiratory tract infection, reproductive system infection, complications of pregnancy, contraceptive intake, folic acid intake, housing renovation and strong tea drinking. The exposure time of maternal variables was defined as from 6 months before conception through the first trimester of pregnancy. Occupational hazards exposure was defined as having been exposed to those toxic and hazardous substances in their workplace, including organic solvents , noxious gases , heavy metals , X-ray, noise, etc. Premarital medical examination was used for couples to get married, in order to prevent diseases that might affect the health of offsprings and promote reproductive health, including the testing of serious hereditary diseases, infectious diseases, and psychiatric disorders. Chronic disease was defined as mothers or fathers had suffered from chronic diseases in 6 months before conception, such as heart disease, kidney disease, liver disease, hypertension, diabetes, anemia, etc. Housing renovation was defined as the house lived by mother had been renovated not more than 6 months. Strong tea drinking was defined as more than 200 ml per day on average. Pickled/smoked food intake, vegetable and fruit intake, fish/shrimp/meat/egg intake, and milk/soymilk intake were classified into three scales (times/week): ≤ 2, 3–5, >5, and the exposure time was defined as the first trimester of pregnancy. Smoking referred to active smoking in the study, and the exposure levels were classified into five scales (cigarettes/day): 0, 1–10, 11–20, 21–40, >40. Alcohol drinking was defined as drinking any liquor, including beer, wine and white spirit in the first trimester of pregnancy, the exposure levels were classified into three scales (times/week): 0, 1–2, ≥3.Characteristics and conditions of the fathers In the present study, there were six variables related to the fathers, including age, occupational hazard exposure, chronic disease, smoking, alcohol drinking, and strong tea drinking. The definitions of paternal variables were the same as the maternal variables, and the exposure time was defined as 6 months before their wives’ conceptions.We modified the questionnaire based on the pilot study. Before the formal survey, unified and strict training was provided to all of the investigators. The subjects were strictly selected according to the inclusion criteria and the diagnosis criteria. Five percent of all of the completed questionnaires were reviewed randomly, and the questionnaires with missing data >10 % and/or errors in logic >10 % were excluded from the study. To ensure the quality of the data entry, dual input was used, and logic checks were performed on the input data.A large number of variables (28 variables) were investigated in this study. We used univariate logistic regression to identify the NSCL/P-associated significant risk factors and then used Fisher discriminant analysis to establish a simple and useful prediction model based on the significant predictors. Univariate analysis could not control the confounding effect of other variables, or avoid the collinearity of some variables. Thus, in the Fisher discriminant analysis, we used a stepwise method to determine the final prediction, which could control the confounding effect and overcome the collinearity between variables.Z: discriminant scores between two groups; X1, X2, X3, ⋯, Xm: discriminant variables; C1, C2, C3, ⋯, Cm : discriminant coefficients for each discriminant variable. The discriminant variables could be selected via two methods: ‘enter variables together’ and ‘enter variables stepwise’. The stepwise method selected the discriminant variables on basis of Wilks’ lambda statistic, and in general, the F value was set at F Entry = 3.84 and F Removal = 2.71. The discriminant function established by stepwise discriminant was simpler and more effective. Assuming that the mean discriminant score of the controls was Zi for each subject; if Zi>Zi≤Fisher discriminant was to find a linear combination for categorical groups, as the discriminant scores (Z) were calculated to maximize the between-group variance and minimize the within-group variation. The linear combination was known as a Fisher discriminant function as follows:P <0.05.Using Epidata 3.1 software , we constructed a database and then entered the data. The data that were obtained were analyzed using SPSS 18.0 software . The results were considered to be significant at A total of 363 subjects who were admitted between July 2012 and June 2013 were surveyed (122 cases and 241 controls), and 24 subjects (9 cases and 15 controls) were excluded from the study because they refused to participate in the study, or the data collected was incomplete. Finally, 339 questionnaires were included in the study , comprising 113 cases  and 226 controls . Table Using univariate logistic regression analysis, 28 variables were analyzed in sequence, including maternal and paternal variables relevant to NSCL/P.Based on the univariate logistic regression analysis, the following 13 variables were significantly associated with NSCL/P Table : low matEntry = 3.84, F Removal = 2.71) based on the screened 13 variables that were statistically significant. The stepwise discriminant analysis showed that Wilks’ lambda, as a test of the discriminant function, was significant , and 8 variables were selected, as follows: family income (X1), maternal occupation hazards exposure (X2), premarital medical examination (X3), housing renovation (X4), milk/soymilk intake in the first trimester of pregnancy (X5), paternal occupational hazards exposure (X6), paternal strong tea drinking (X7), and the family history of NSCL/P (X8). The final standardized discriminant function was calculated according to the following Equation:Using the results of the univariate logistic regression analysis, a risk prediction model of NSCL/P was constructed by a stepwise Fisher discriminant analysis  curve. The area under the ROC curve (AUC) is typically between 0.5 and 1.0. When the AUC is between 0.5 and 0.7, the diagnostic value of the test is low; when it is between 0.7 and 0.9, it has a medium diagnostic value; and when it is more than 0.9, it has high diagnostic value.P < 0.001, 95 % CI: 0.794~0.898). The diagnostic value of the model was medium.The AUC of the discriminant analysis prediction model is shown in Fig. NSCL/P is a common congenital anomaly, which seriously affects children’s health. The etiology of NSCL/P is complex and largely unknown. Recently, most studies have focused on the identification of risk factors of NSCL/P, while an effective risk prediction tool for NSCL/P is lacking. In the present study, the prediction model established by discriminant analysis was successful in classifying 83.8 % of participants, with an AUC of 0.846. The prediction model can be used as a risk prediction tool for NSCL/P, as it aims to identify the high-risk population of NSCL/P in the first trimester of pregnancy and to provide important information for a further clinical ultrasound in the second or third trimesters of pregnancy. The pregnant women with a high predictive risk were identified as the population at a high risk of NSCL/P and listed as a focus group for clinical prenatal ultrasound diagnosis. In addition, the prediction model also can be applied by doctors into pre-conception counseling and education for women of childbearing age. If women of childbearing age discover that they are at a high risk by this prediction, they may be able to control some important risk factors to reduce the risk of NSCL/P during pregnancy. To the best of our knowledge, there was no available information on predicting the occurrence of NSCL/P. Accordingly, this is the first study using a discriminant analysis to predict the risk of NSCL/P in pregnant women.In the present study, 13 factors screened by univariate logistic analysis were related to NSCL/P, but only 8 factors used as predictor entered the discriminant function. Consistent with previous studies, a low family income , not attIn this study, the NSCL/P risk prediction model had good specificity, while the sensitivity was not satisfactory. The sensitivities (the rates of correct prediction for the NSCL/P cases) and the specificities (the rates of correct prediction for the controls) were 74.3 and 88.5 %, respectively. There were two reasons for the low sensitivity. First, the 8 predictors selected by the discriminant analysis except for the family history were common risk factors of congenital anomalies but were not specific indicators for NSCL/P. Second, due to the small sample size and the low exposure rates of some of the investigated factors, some common important risk factors were not included in the prediction model, such as maternal age, folic acid intake, history of infection during pregnancy, mothers’ abnormal reproductive history, medication use during pregnancy, maternal stressful events during pregnancy, tobacco, and alcohol. Many of the published papers show conflicting results on the relationship between maternal age and NSCL/P , 12, 41.The present study has specific limitations. First, we used case–control data to select the predictors, and this inevitably led to recall bias in the data. Second, because of the limitations of the sample size, a self-verification was adopted to evaluate the discriminant predictive effect of NSCL/P, which tended to exaggerate the discriminant effect. Further studies are needed to confirm the validity and reliability of the NSCL/P prediction model in the larger population. Third, the 95 % confidence intervals (CI) of odds ratios (OR) for some of the screened factors  were wide due to the small sample size. The corresponding ORs were significant, but with limited precision and reliability. Finally, due to the low exposure rates of some of the investigated factors, certain important risk factors of NSCL/P failed to enter the prediction model, resulting in its low sensitivity. We will need to conduct additional research to identify the specific predictors of NSCL/P to improve the sensitivity and specificity of the model and attempt to construct the prediction model by other statistical methods, such as artificial neural networks, decision trees or logistic regression, to modify and improve the prediction model.The discriminant prediction model, which is based on family income, maternal occupational hazards exposure, premarital medical examination, housing renovation, drinking milk/soymilk in the first trimester of pregnancy, paternal occupational hazards exposure, paternal drinking of strong tea, and family history of NSCL/P, is useful for predicting the risk of NSCL/P. Further research is needed to improve the model, and confirm the validity and reliability of the model.


Additionally, this association was also observed in cohort studies  and case-control studies . However, When the intrahepatic cholangiocarcinoma and extrahepatic cholangiocarcinoma were analyzed separately, the present study only indicated cholecystectomy was associated with increased the risk of extrahepatic cholangiocarcinoma , rather than intrahepatic cholangiocarcinoma . In conclusion, cholecystectomy was associated with a significant 54% increase in the risk of cholangiocarcinoma, especially in the extrahepatic cholangiocarcinoma.Studies have reported that cholecystectomy may increase the risk of cholangiocarcinoma. However, this association is controversial. Thus, we conducted a systematic review and meta-analysis to explore the relationship between cholecystectomy and the risk of cholangiocarcinoma. Relevant studies were identified by searching PubMed, EMBASE, ISI Web of Science published before February 2017. We used the random effects model proposed by DerSimonian and Laird to quantify the relationship between cholecystectomy and risk of cholangiocarcinoma. Publication bias was evaluated using funnel plots, Begg's and Egger's tests. Subgroup and sensitivity analyses were performed to validate the stability of the results. 16 articles, comprising 220,376 patients with cholecystectomy and 562,392 healthy controls, were included in our research. Our meta-analysis suggested that the risk of cholangiocarcinoma was significantly higher in the cholecystectomized patients in comparison with healthy controls, with heterogeneity among studies (summary odds ratio [OR] = 0.72; confidence interval [CI] = 0.55–0.90; I Gallstones are abnormal masses of a solid mixture of cholesterol crystals, mucin, calcium bilirubinate, and proteins . GallstoCholangiocarcinoma, which was first described by Durand-Fardel in 1840, is a malignant tumor originating from bile duct epithelium . CholangFigure The main characteristics of the included studies are listed in Table 2 = 69.5%; p = 0.006) , rather than ICC  , with significant heterogeneity   to evaluate the effect of cholecystectomy on the risk of cholangiocarcinoma. Therefore, the findings may provide us insight into the relationship between cholecystectomy and the risk of cholangiocarcinoma, and these results are of potential interest to the field of cholangiocarcinoma research. Secondly, subgroup and sensitivity analyses were performed to determine the factors that may affect results. It makes our findings more reliable. Third, we performed a comprehensive literature search of the PubMed, EMBASE and Web of Science databases to identify potential studies to investigate the relationships between cholecystectomy and the risk of cholangiocarcinoma. In addition, most of the studies included in our meta-analysis were of high quality. All of these characteristics make the conclusions of our study more convincing.There are several limitations that should be considered. First, most of the studies included in our meta-analysis were case-control studies, which was prone to generate recall and selection biases. Additionally, the heterogeneity among studies was significant because of different study designs and demographic characteristics inconsistency. Second, the present study only investigated the risk of cholangiocarcinoma in patients with cholecystectomy compared with healthy population. As a result of the restricted number of included studies in the analysis, the risk of cholangiocarcinoma in patients with cholecystectomy compared with gallstone patients was not explored. Third, what is being observed is just an association, which is subject to confounding bias. The established risk factors for cholangiocarcinoma include primary sclerosing cholangitis, hepatolithiasis, bile-duct cysts and parasitic infections . HoweverIn summary, our meta-analysis indicated that the risk of cholangiocarcinoma was associated with a 54% increase in patients who had cholecystectomies in comparison with healthy controls, and the relationship was also demonstrated in cohort and case-control studies. However, When the ICC and ECC were analyzed separately, the present study only indicated cholecystectomy was associated with increased the risk of ECC, rather than ICC. More prospective studies and basic research are still needed to validate the association of cholecystectomy and cholangiocarcinoma risk and the potential mechanisms.We searched published reports in the PubMed, EMBASE and Web of Science databases using the following keywords:  and (“biliary tract cancer” OR “bile duct cancer” OR “biliary tract neoplasms” OR “cholangiocarcinoma”). We placed no restrictions on the language or date of publication.The eligibility criteria were as follows: study design (case control or cohort); cholecystectomy as the exposure factor and cholangiocarcinoma or bile duct cancer or biliary tract cancer as the outcome; and odds ratio (OR)/risk ratio (RR) values and corresponding 95% confidence intervals available or sufficient information to calculate them. If two studies reported the same data, we selected the study with the larger sample.Two researchers (Y.W. and A.W.) independently extracted the required information from the selected studies in a standardized manner. We collected the following information from each article: first author's name, year of publication, country of origin, study design (case-control or cohort), number of participants, duration of follow-up, sources of controls, adjustment for confounding variables, and OR/RR values and 95% CIs.The Newcastle-Ottawa Scale (NOS)  was usedThe OR/RR values and corresponding 95% CIs were used to evaluate the risk of cholangiocarcinoma in with a history of cholecystectomy. We treated hazard ratios as equivalent to RRs. We used the random effects model proposed by DerSimonian and Laird to quantify the relationship between cholecystectomy and the risk of cholangiocarcinoma .2 statistic was used to quantify the heterogeneity between studies, and I2 values of 25%, 50%, and 75% represented low, medium, and high heterogeneity, respectively [P values less than 0.1 indicated that clear heterogeneity existed. Publication bias was qualified with funnel plot and Begg's [P values less than 0.05 indicated the presence of bias.The Iectively . P valued Begg's  and Egged Begg's  tests, aWe also performed subgroup analyses by subtype of cancer, geographic areas, study design and whether liver fluke infestation, cholangitis, gallstones, alcohol intake or smoking were adjusted for in the models. Sensitivity analysis was conducted to assess the stability of the results by sequentially excluding one study in one turn. Additionally, sensitivity analyses were also performed by changing the pooling model (random-effects model or fixed-effects model) and excluded studies that the NOS sources were < 7.All statistical analyses were performed using STATA version 12.0 (Stata).


The association between fibroblast growth factor receptor 2 (FGFR2) polymorphism and breast cancer (BC) susceptibility remains inconclusive. The purpose of this systematic review was to evaluate the relationship between FGFR2  polymorphism and BC risk. PubMed, Web of science and the Cochrane Library databases were searched before October 11, 2015 to identify relevant studies. Odds ratios (ORs) and 95% confidence intervals (CIs) were used to estimate the strength of associations. Sensitivity and subgroup analyses were conducted. Thirty-five studies published from 2007 to 2015 were included in this meta-analysis. The pooled results showed that there was significant association between all the 3 variants and BC risk in any genetic model. Subgroup analysis was performed on rs2981582 and rs2420946 by ethnicity and Source of controls, the effects remained in Asians, Caucasians, population-based and hospital-based groups. We did not carryout subgroup analysis on rs2981578 for the variant included only 3 articles. This meta-analysis of case-control studies provides strong evidence that FGFR2  polymorphisms were significantly associated with the BC risk. For rs2981582 and rs2420946, the association remained significant in Asians, Caucasians, general populations and hospital populations. However, further large scale multicenter epidemiological studies are warranted to confirm this finding and the molecular mechanism for the association need to be elucidated further. Breast cancer (BC), one of the most common malignant tumors among women worldwide, has the highest mortality rate in female cancer. Its incidence rate is increasing year by year and the patients are becoming younger and younger in the world , 2. BC iIn recent years, genome-wide association study (GWAS) provides a good technical support for the study on the susceptibility loci with high variation frequency and low penetrance . Large nLots of researches have reported the association between FGFR2  polymorphism and BC risk. However, due to differences in ethnic and regional and other factors, the conclusions of related reports are still inconclusive. Raskin et al  found thThe specific search process is shown in Figure The FGFR2  polymorphisms genotype distribution and allele frequencies incase groups and control groups were shown in Table For rs2420946, 11 studies with 7,840 cases and 9,349 controls were included to assess the association. As shown in Table 3 papers with 833 cases and 1012 controls were adopted to evaluate the association between the rs2981578 polymorphism and the BC risk. As shown in Table As shown in Table Q statistic. Random-effect models were applied if p-value of heterogeneity tests were less than 0.1 (p ≤ 0.1), otherwise, fixed-effect models were selected  polymorphism and BC risk. However, due to differences in ethnic and regional and other factors, the conclusions of related reports are still inconclusive. Thus, we conducted the meta-analysis to evaluate the relationship between FGFR2  polymorphism and BC risk.In our study, there were 31 studies with 54,677 cases and 80,418 controls for FGFR2 rs2981582 variants. In the total population, the pooled results indicated that the correlation between FGFR2 rs2981582 polymorphism and the occurrence of BC was significant in any genetic model. Furthermore, in ethnicity-specific analysis, FGFR2 rs2981582 were also significantly associated with BC risk both in Asians and Caucasians. We didn't discuss the African subgroup for just one study was from African. The analysis in different source of controls showed the same association between FGFR2 rs2981582 polymorphism and BC susceptibility both in HB and PB, indicating that both hospital populations and general populations followed the same relationship. For rs2420946, 11 studies with 7,840 cases and 9,349 controls were included to assess the association. In the total population, the pooled ORs suggested that rs2420946 was significantly associated with BC susceptibility in all the five genetic models. When stratified by ethnicity and source of controls, the results showed the same association in Asians, Caucasians, hospital populations and general populations, indicating that different genetic backgrounds and living environment were not strong enough to change these associations. All the results for the two variants  were partially consistent with the consequences of Wang's , Peng's Three articles with 833 cases and 1012 controls were adopted to evaluate the association between the rs2981578 polymorphism and the BC risk. As the preceding two variants, the association between rs2981578 variant and BC susceptibility was also significant in any genetic model. For just only 3 studies, no stratified study was conducted for rs2981578 polymorphism. However, in Zhou's  meta-anaOur meta-analysis has several limitations. First, our study is a summary of the data. For lack of all individual raw data, we could not assess the cancer risk stratified by other covariates including age, sex, environment, hormone level, menopause age and other risk factors. We also cannot analyze the potential interaction of gene-environment and gene-gene. Second, only published papers were included in our meta-analysis, there still may be some unpublished studies which are in line with the conditions. Therefore, publication bias may exist even no statistical evidence was found in the meta-analysis. Third, for just only 3 papers, no stratified study was conducted for rs2981578 polymorphism. Moreover, our study is a summary of the data. We did not verify it from the level of molecular mechanism. Data from large scale multicenter epidemiological studies is still needed to confirm the relationship between FGFR2  polymorphisms and BC risk, and the molecular mechanism for the associations need to be elucidated further.In conclusion, our meta-analysis based on case-control studies provides strong evidence that FGFR2  polymorphisms are significantly associated with the BC risk. For rs2981582 and rs2420946, the association remained significant in Asians, Caucasians, general populations and hospital populations. However, further large scale multicenter epidemiological studies are warranted to confirm this finding and the molecular mechanism for the associations need to be elucidated further.We searched PubMed, Web of science and the Cochrane Library for relevant studies published before October 11, 2015. The following keywords were used: (Fibroblast Growth Factor Receptor 2 or FGFR2) and (variant* or genotype or polymorphism or SNP) and (breast) and (cancer or carcinom* or neoplasm* or tumor), and the combined phrases for all genetic studies on the association between the FGFR2  polymorphism and BC risk. The reference lists of all articles were also manually screened for potential studies. Abstracts and citations were screened by two researchers independently. All the eligible articles need a second screening for the full-text. The searching was done without language limitations.i)independent case-control studies for humans; ii) the study evaluating the association between FGFR2  polymorphism and BC risk; iii) the study presenting available genotype frequencies in cancer cases and control subjects for risk estimated; iiii) cases should have been diagnosed by a pathological examination. We excluded comments, editorials, systematic reviews and studies lacking sufficient data or studies with male cases. If the researches were duplicated or shared in more than one study, the most recent publications were included.Inclusion criteria: A study was included in this meta-analysis if it met the following criteria: We used endnote bibliographic software to construct an electronic library of citations identified in the literature search. All the PubMed, Web of science and the Cochrane Library searches were performed using Endnote. Duplicates were found automatically by endnote and deleted manually. All data extraction were checked and calculated twice according to the inclusion criteria listed above by two independent investigators. Data extracted from the included studies were as follows: First author, year of publication, country, Ethnicity, Source of controls, Genotyping method, number of cases and controls and evidence of Hardy-Weinberg equilibrium(HWE) in controls. A third reviewer would participate if some disagreements were emerged, and a final decision was made by the majority of the votes.2χ test in the control group of each study [Z test was used to assess the significance of the ORs, 2I and Q statistics was used to determine the statistical heterogeneity among studies. A random-effect model was used if p value of heterogeneity tests was no more than 0.1 (p ≤ 0.1), and otherwise, a fixed-effect model was selected [p< 0.05 was considered statistically significant. We performed subgroup according to Ethnicity and Source of controls.All statistical analyses were performed using STATA version 11.0 software  and Review Manage version 5.2.0 . Hardy-Weinberg equilibrium (HWE) was assessed by ch study . The strselected , 22. Senselected , 24. The