CYP4A11 and essential hypertension, though with conflicting results. To elucidate this relationship, a case-control study and meta-analysis were performed to assess the possible association of essential hypertension with CYP4A11 genetic variations.CYP4A11 oxidizes endogenous arachidonic acid to 20-hydroxyeicosatetraenoic acid, a renal vasoconstrictor and natriuretic in humans. Previous studies demonstrated an association between a functional variant (T8590C) of CYP4A11 variant. To further investigate the association, we conducted a meta-analysis including eight studies published previously in July 2012.Associations between the T8590C polymorphism and essential hypertension were examined in 328 unrelated cases and 297 age-matched controls in Han Chinese individuals. High-resolution melting was used to identify the CYP4A11 T8590C polymorphism showed no significant difference between cases and controls . However, the meta-analysis showed that the CYP4A11 T8590C polymorphism may increase the risk of essential hypertension in an additive model , a dominant model , a recessive model  and a homozygote contrast . Also, a significant relationship was observed among Caucasians in the additive model, the homozygote contrast, the recessive model and the dominant model . However, no association was observed in an Asian population .The frequency of the CYP4A11 T8590C variant and essential hypertension, especially in Caucasians. The case-control study did not find a significant association among the Han Chinese population, but the controls were poorly matched and meaningful conclusions cannot therefore be made. Further large-scale studies are needed to clarify whether the CYP4A11 T8590C polymorphism is associated with hypertension risk in Asians or has a gender-specific effect.This meta-analysis suggests there is a significant association between the  Hypertension is a major risk factor for cardiac, renal and cerebrovascular morbidity and mortality Hypertension has a substantial heritability that is likely to be of polygenic origin. To date, several variants have been identified by genome-wide association studies (GWASs) to be associated with essential hypertension Cytochrome P450 (CYP) is a superfamily of cysteine-heme enzymes that are key mediators of the oxidative transformation of endogenous and exogenous molecules. Cytochrome P450 enzymes are classified into families, subfamilies and individual isoenzymes based on similarities in their amino acid sequences. Human CYP4A11, which belongs to the cytochrome P450 family 4, catalyzes the conversion of endogenous arachidonic acid to 20-tetrahydrocannabinol acid (20-HETE) CYP4A11 gene is located on chromosome 1p33, spans approximately 12.57 kilobase-pairs and contains 12 exons. More than 120 single nucleotide polymorphisms (SNPs) of the human CYP4A11 gene are listed in the National Center for Biotechnology Information SNP database (http://www.bianca.nl.NIH.gov/SNP). One of the coding SNPs, rs1126742 (T8590C), which results in a phenylalanine to serine substitution at amino acid 434 in exon 11 of CYP4A11, has been found to be associated with blood pressure (BP) in several populations CYP4A11 T8590C variant has a role in modulating BP in the Chinese Han population remains unknown. Given this background, the aim of this study was to investigate the possible association between the CYP4A11 T8590C variant and essential hypertension in the Chinese Han population. In addition, a meta-analysis was conducted, including studies examining the association between the CYP4A11 T8590C variant and the risk of hypertension, to clarify previous inconsistencies.Given the role of 20-HETE in the regulation of hypertension, CYP4A11 is the subject of much research interest. The human Patients diagnosed with essential hypertension were recruited at Guangzhou General Hospital of Guangzhou Military Command from 2008 to 2011. Three hundred and twenty-eight essential hypertension patients and 297 age-matched controls were enrolled in the study. BP was measured twice in each participant while seated, after 5 min of rest. Essential hypertension was diagnosed based on the following criteria: seated systolic BP greater than 140 mmHg or diastolic BP greater than 90 mmHg on three occasions within 2 months after the first BP reading, or current use of antihypertensive agents. Individuals with secondary hypertension, cancer, chronic kidney or hepatic disease, stroke, diabetes mellitus or severe psychiatric illness were excluded. All individuals enrolled were from the Han population in China. Diagnosis of diabetes was based on the history of treatment for hypoglycemia and/or confirmed fasting blood glucose >7.0 mmol/l. Individuals who had smoked ≥10 cigarettes per day for at least 2 years were defined as smokers and those who had drunk ≥210 g alcohol per week for more than 3 years were defined as drinkers.This study protocol was approved by the Ethics Committee of Guangzhou General Hospital of Guangzhou Military Command. All individuals in this manuscript gave written informed consent for the publication of their case details.CYP4A11 T8590C allele was genotyped by high resolution melting. The primers used were as follows: forward: 5′-TGGGTGGCAAG TAGGTGCTGGA-3′; reverse: 5′-TGTTGAGCAGAACCCGGTGCA-3′. PCR reactions were performed in a final volume of 20 µl using Highlighter 480 High Resolution Melting Master  and 30 ng DNA. The primers and MgCl2 were used at 10 µM and 2.5 M, respectively. The PCR program comprised an initial denaturation-activation step at 95°C for 10 min, followed by a 45-cycle program of denaturation at 95°C for 10 s, annealing at 57.8°C for 10 s and elongation at 72°C for 15 s with reading of the fluorescence and a single acquisition mode. The melting program comprised three steps: denaturation at 95°C for 1 min, renaturation at 40°C for 1 min and then melting, with a continuous fluorescence reading from 60 to 90°C at 25 acquisitions per °C.Peripheral venous blood was drawn from each individual. Genomic DNA was extracted from whole blood using the Tiangen DNA blood kit . The t-test. Differences of continuous variables departing from the normal distribution, even after transformation, were tested by the Mann-Whitney U-test. Odds ratios (ORs) and their 95% confidence intervals (CIs) were calculated to determine associations between the polymorphisms and essential hypertension. Two-tailed P values <0.05 were considered statistically significant.Statistical analysis was performed using the SPSS 13.0 software package . The genotype frequency distribution for each variant was separately tested for Hardy-Weinberg equilibrium (HWE) with a chi-square test in the case and control groups. Differences of continuous variables with a normal distribution (presented as mean ± SD) between the two groups were calculated using the independent-sample CYP4A11 and essential hypertension in different studies and ethnicities, we performed a meta-analysis of previously published studies  concerning the T8590C polymorphism and essential hypertension. Our own data were also included. We systematically searched PubMed, Medline, Embase and the China National Knowledge Infrastructure (CNKI) using the following keywords: “CYP4A11”, “cytochrome P450 4A11”, “polymorphism”, “gene” and “hypertension”. We also retrieved additional studies by hand, searching the bibliographies of original research reports and review articles, and through the Medline option “related articles”. All relevant referenced articles were also investigated.To explore the association between the T8590C polymorphism of CYP4A11 T8590C polymorphism and the risk of developing hypertension; (b) unrelated case-control studies or cohort studies; (c) availability of the sample size, ORs and 95% CIs or information including genotype frequency of cases and controls that could help in inferring the study results; and (d) presence of non-overlapping data. For duplicate publications, the study with the smaller data set was excluded. Diagnoses of essential hypertension were made in accordance with the diagnostic requirements of the World Health Organization established in 1999. Subjects with systolic BP≥140 mmHg or diastolic BP≥90 mmHg and subjects who had no secondary hypertension were included.The inclusion criteria were: (a) evaluation of the Two investigators reviewed and extracted data from all of the eligible publications independently, according to the inclusion and exclusion criteria listed above. All data were collected according to a standard protocol. Studies that were repeated, of poor research quality, did not meet the inclusion criteria or provided little information or insufficient data were excluded. Study quality was assessed using the 9-star Newcastle-Ottawa Scale CYP4A11 T8590C polymorphism and essential hypertension was compared between the hypertension and control groups by the OR and the corresponding 95% CI. We evaluated the risk using an additive model , a dominant model (TC+CC versus TT), a recessive model (CC versus TT+TC) and the homozygote contrast (comparison of CC versus TT).The association between the Q-test, and was considered statistically significant when P<0.10. The inconsistency index I2 was also calculated to evaluate the variation caused by the heterogeneity. A high value of I2 indicated a higher probability of the presence of heterogeneity . If there was heterogeneity among the studies (P<0.1), a random-effects model was used to estimate the combined OR. Otherwise, the pooled OR was estimated by the fixed-effect model Z-test and the significance was set at P<0.05. Fisher’s exact test was used to assess the HWE and the significance was set at P<0.05. In addition, sensitivity analysis was carried out by including and excluding studies not in HWE. Potential publication bias was estimated using a funnel plot. Egger’s linear regression test was used to evaluate the funnel plot asymmetry on the natural logarithmic scale of the OR . Review Manager 5.1 and STATA 11.0 software were used to perform statistical analysis .The heterogeneity of the studies was tested using the chi-square-based P<0.001) and low-density lipoprotein cholesterol (P = 0.005) than the controls. Additionally, there were significantly higher numbers of males (P = 0.041) and smokers (P<0.001) in the cases compared to the controls.Baseline characteristics of the study population are shown in P = 0.636), as shown in P = 0.477), the recessive model (P = 0.418) and the additive model (P = 0.371). The control group was poorly matched to the cases, particularly with regards to other known predictors of hypertension (gender and smoking), which make the results here difficult to interpret.For the total population, the frequencies of the 8590TT, 8590TC and 8590CC genotypes were 52.0%, 40.8% and 7.2%, respectively. These frequencies did not deviate from those predicted by HWE. The allele frequencies of the 8590C allele in cases and controls were 26.5% and 28.8%, respectively. No significant difference in the genotype frequency was observed between the cases and controls , the dominant model , the recessive model  and the homozygote model . Seven studies also underwent stratified analysis by ethnicity (one study was excluded because of mixed ethnicity). No increased risk was detected among Asians in the additive model, the homozygote contrast, the recessive model or the dominant model . However, significant differences were detected among Caucasians in the additive model, the homozygote contrast, the recessive model and the dominant model, as shown in P<0.05). Furthermore, four studies that conducted subgroup analysis by gender were pooled and analyzed in subgroups. No statistically significant differences were found for male or female subjects in any of the various statistical models , as shown in The findings for heterogeneity and the combined association between the I2 = 59%, Pheterogeneity = 0.03). Then, random-effects models were used to evaluate combined ORs when necessary. Sensitivity analysis was performed both by sequential removal  of individual studies and by cumulative statistical analysis comparing subjects worldwide.Significant heterogeneity between studies was observed in some comparisons; detailed data are shown in P values for the Q-tests became significant (P<0.05) after deletion of the study by Fava et al. Among female subjects, the combined ORs of the additive model, the recessive model and the homozygote contrast were influenced by the present study. The P values for the Q-tests became significant (P<0.05) after deletion of this study. Other combined ORs for the CYP4A11 T8590C polymorphism were not influenced by the addition or removal of any individual study.For the total population, sensitivity analyses indicated that the present study was the main source of heterogeneity. In addition, among male subjects, the combined ORs of the additive model and the dominant model were influenced by one study in HWE. The T = 1.99, P = 0.09). The funnel plot and Egger’s test suggested that the selection of publications was an unlikely source of bias in this meta-analysis of the association between the CYP4A11 T8590C polymorphism and essential hypertension.A funnel plot and Egger’s test were performed to assess publication bias in this meta-analysis. The funnel plot showed no apparent evidence of publication bias . There wCYP4A11 has been reported to encode an enzyme that converts arachidonic acid to 20-HETE, primarily in the kidney CYP4A11 gene is a candidate causative gene for hypertension CYP4A11 and essential hypertension. CYP4A11 is known to be highly polymorphic, expanding the possibilities for studying the mechanism of action of this gene, especially as a risk factor for cardiovascular diseases et al. reported recently that CYP4A11 is highly expressed in human renal tubules and regulates 20-HETE, which is associated with the function of renal tubules and salt-sensitive hypertension et al. reported that the 8590T/C polymorphism of CYP4A11 was functional, significantly reducing the catalytic activity of CYP4A11 enzyme and affecting the synthesis of 20-HETE et al. were also the first to demonstrate a positive association between 8590T/C and hypertension in whites as a loss-of-function variant. However, in 120 black subjects from the Tennessee cohort, there was no association between the 8590C allele and hypertension CYP4A11 T8590C polymorphism and hypertension has also been studied by Mayer et al. et al. also reported that carriers of the C allele had a higher prevalence of hypertension in a prospective cohort including 6002 Swedish subjects CYP4A11 gene in Australians CYP4A11 8590T/C polymorphism and hypertension was observed by Sugimoto et al. in Japanese subjects The product 20-HETE, the monooxygenase derivative of arachidonic acid synthesized by CYP450 enzymes, is a candidate for the regulation of BP and hypertension. In humans, two members of the CYP4A subfamily have been detected in the kidney. P = 0.636) in the total population  CYP4A11 T8590C polymorphism appears to be a useful genetic marker of essential hypertension in men P values became significant.It is well established that the prevalence of hypertension and cardiovascular disease differs between men and women, the latter being protected from cardiovascular events until menopause; there is a rapid increase in the risk profile after menopause. Gene-specific effects have been indicated as responsible, but most genetic studies do not include sex as a variable in data analysis, and a recent meta-analysis pooling thousands of subjects of previous GWASs did not identify any sex-specific effect of genes associated with BP/hypertension. Experimental models suggest that enzymes involved in 20-HETE biosynthesis and metabolism have a role in the control of BP with a sex-specific effect CYP4A11 T8590C polymorphism is associated with hypertension risk in Asians or has a gender-specific effect.When interpreting these results, caution should be applied. Relatively large heterogeneity was present in the meta-analysis, perhaps due to the limited study sample size. A relatively limited number of studies and samples were analyzed, partly because the original studies had rather limited sample sets. Moreover, differences in the ethnicity and selection of controls, age distribution and lifestyle factors may have contributed to heterogeneity. Further large scale studies are needed to clarify whether the CYP4A11 T8590C polymorphism is associated with risk of hypertension in the general population, especially in Caucasians.In summary, our meta-analysis suggests that the Figure S1PRISMA flow diagram of the process of identifying and including articles for the systematic review.(DOC)Click here for additional data file.Checklist S1PRISMA checklist.(DOC)Click here for additional data file.


The association of p53 codon72 polymorphism with prostate cancer has been widely reported; however, the results are inconsistent. To derive a more precise estimation of this relationship, we performed an updated meta-analysis from 10 case-control studies. We conducted a search in the PubMed database without a language limitation, covering all papers published until July 2010. Risk ratios (RR) with 95% confidence intervals(CIs) were used to assess the strength of the association. Ten studies including 1,196 cases and 1,704 controls were selected. Overall, no significant differences of total prostate cancer risk and p53 codon polymorphism was found . In the stratified analysis by ethnicity, the same results were found. However, in the control subgroup, there was a modest decreased association between prostate cancer risk and population-based control subjects under the recessive genetic model . This meta-analysis suggested that p53 codon Pro72Arg polymorphism could be weakly associated with prostate cancer risk.The tumor suppressor gene  Prostate cancer (PCa) is the most commonly seen male malignancy and the second leading cause of cancer deaths in men in the United States, with estimated 192,000 new cases and 27,000 deaths in 2009TP53 gene, which is located on chromosome 17q13p53 gene and its encoded protein play a central role in regulating cell cycle progression, DNA repair, cellular growth and apoptosisThe p53 transcription factor is encoded by the A common variant, a single nucleotide polymorphism (SNP) at codon72 , alters activities of p53. The C to G change results in a proline (Pro) to arginine (Arg) amino acid substitution in the proline-rich region that is essential for p53-mediated apoptosisp53 gene codon72 polymorphism and tumor formation has been extensively studied, including in ovarian, lung, cervical, and colon cancer and PCa. Of the ten publications on PCap53 codon Pro72 polymorphism was associated with a decreased risk of PCa, while others reported no association or an association between Arg and decreased risk of PCa. Therefore, whether the polymorphism of p53 gene codon72 is associated with PCa or not is still controversial.The association between p53 codon72 in PCa, and to derive a more precise estimation of the association of Pro72Arg polymorphism in p53 and PCa, we performed a meta-analysis of all eligible case-control studies.Taking into consideration the extensive role of http://www.ncbi.nlm.nih.gov/), last search updated on July 2010, with the keywords “p53” or “TP53”, “polymorphism” and “prostate cancer” or “prostate”. Using these terms, a total of 101 articles were retrieved, of which 10 articlesp53 codon Pro72Arg polymorphism and PCa risk.We conducted searches on the PubMed database  the study assessed the correlation between global cancer and at least one of the polymorphisms cited above; 2) case-control studies; 3) control subjects were matched with case patients in age and gender; 4) only full-text manuscripts were included. Major exclusion criteria were: 1) no control population; 2) no available genotype frequency; 3) duplication of previous publications; 4) manuscripts with a clear bias of accrual.Studies testing the association between Two of the authors reviewed the results of each of the database searches to make sure that all published papers were not missed. Data were collected based on the first author's last name, year of publication, country of origin, ethnicity, cancer type, sample size (cases/controls), genotyping methods, age range in cases and controls, source of control and Hardy–Weinberg equilibrium (HWE) of controls.p53 codon Pro72Arg gene was conducted using polymerase chain reaction-restriction fragment length polymorphism (PCR-RFLP)Genotyping for SNP of p53 codon Pro72Arg polymorphism and PCa based on the genotype frequencies in cases and controls. Subgroup analysis stratified by ethnicity was performed first. Ethnicity was categorized as Caucasian, Asian and African. Source of control subgroup analysis was performed on two classifications: population-based and hospital-based.Crude risk ratios (RR) with 95% confidence intervals (CIs) were used to measure the strength of the association between Z test. Heterogeneity assumption was evaluated with a chi-square-based q test among the studies. A P value of more than 0.05 for the q-test indicated a lack of heterogeneity among the studies. In order to better evaluate the extent of heterogeneity between studies, the I2 test was also used. As a guide, I2 values of <25% may be considered ‘low’, value of ∼50% may be considered ‘moderate’ and values of >75% may be considered ‘high’P≤0.05, or I2≥50%, a random-effects model using the DerSimonian–Laird methodP > 0.05, and I2<50%, a fixed-effects model using the Mantel–Haenszel methodp53 codon Pro72Arg, we investigated the association between genetic variants and PCa risk in allelic contrast , homozygote comparison (Pro/Pro vs Arg/Arg), heterozygote comparison (Pro/Arg vs Arg/Arg), dominant genetic model (Pro/Pro+Pro/Arg vs Arg/Arg) and recessive genetic model (Pro/Pro vs Pro/Arg+Arg/Arg). The funnel plot asymmetry was assessed with Egger's test. Publication bias was assessed with Egger's test; P < 0.05 was considered statistically significantp53 codon Pro72Arg polymorphism from expectation under HWE was assessed by X2 test in controls using the Pearson chi-square test for goodness of fit, P < 0.05 was considered significant. All statistical tests for this meta-analysis were performed with STATA software .The fixed effects model and the random effects model were used to calculate the pooled RR. The statistical significance of the summary RR was determined by p53 codon Pro72Arg, and one studyOf the 101 abstracts retrieved through the search criteria, 71 were irrelevant, six articles were reviews, 13 studies were excluded because they did not concern The characteristics of the eligible studies are presented in Pheterogeneity = 0.016), recessive genetic model (Pheterogeneity =0.018) and heterozygote comparison (Pheterogeneity = 0.035), but not for the allelic contrast (Pheterogeneity = 0.084) and the dominant model (Pheterogeneity = 0.077), because the P values were more than 0.05 for Q-tests. However, subgroup analysis regarding ethnicity and source of control were conducted, and the P value for heterogeneity indicated a reduced or absent heterogeneity.As shown in p53 codon Pro72Arg, the results of the meta-analysis are presented in detail in RR = 1.02, 95%CI (0.96-1.09), P = 0.532, I2 = 41.1%; homozygote comparison, RR = 1.12, 95%CI (0.74-1.70), P = 0.590, I2 = 55.8%; heterozygote comparison, RR = 1.22, 95%CI (0.94-1.60), P = 0.136, I2 = 50.1%; dominant models, RR = 1.05, 95%CI (1.00-1.11), P = 0.069, I2 = 51.1% and recessive genetic model, RR = 0.96, 95%CI (0.67-1.37), P = 0.815, I2 = 55.1%]. Additionally, in the stratified analysis by ethnicity, no significant association between PCa and p53 codon72 polymorphism was found in each of the three ethnicities. However, in the subgroup of source of control, we found a significantly decreased risk for PCa in population-based control subjects  than hospital-based. In our results, we found a significantly decreased risk for PCa in population-based control subjects, although the number was quite small and just included two case-control studies. As a result, if the Pro to Arg amino acid alteration occured in the proline-rich region essential for p53-mediated apoptosis, the incidence of PCa would be expected to go down. This maybe helpful for finding the etiology of PCa.p53 codon Pro72Arg polymorphism studiesSeveral limitations in this meta-analysis should be mentioned. First of all, the number of published studies included in our meta-analysis was not sufficiently large for a comprehensive analysis, particularly for any given ethnicity  site. Second, publication bias might have occurred and our Egger's test results may have a substantial risk of being affected by such bias, although the funnel plots as well as Egger's linear regression tests indicated no remarkable publication biases in the meta-analyses. Third, the interactions between gene–gene, gene–environment and even different polymorphic loci of the same gene may modulate PCa risk. Fourth, in some p53 codon Pro72Arg polymorphism was associated with decreased PCa risk in population-based subjects in recessive genetic model. However, no significant association was found in any genetic model in the whole population and ethnic group. Therefore, further well designed large studies, particularly referring to gene-gene and gene-environment interactions, are warranted. These future studies should lead to better and comprehensive understanding of the association between the p53 codon Pro72Arg polymorphism and PCa risk.In summary, our meta-analysis showed the evidence that the


To clarify this conflict, we updated the previously performed meta-analysis by Peck PubMed and Embase databases (through Dec 2012) were searched for English articles on the relationship of the I/D polymorphism in ACE with PICH in humans. Summary odds ratios (ORs) were estimated and potential sources of heterogeneity and bias were explored.A total of 805 PICH cases and 1641 control cases obtained from 8 case-control studies were included. The results suggest that in dominant genetic models, the ACE I/D polymorphic variant was associated with a 58% increase in susceptibility risk of PICH . However, in the subgroup analysis based on race, a significant increased risk was found in Asian DD homozygote carriers (OR = 1.76 and 95% CI = 1.16–2.66 for DD vs. DI+II), but not in Caucasian DD homozygote carriers . The heterogeneity between studies was remarkable, and its major sources of heterogeneity were due to the year in which the study was published. No potential publication bias was observed in dominant genetic models.These data demonstrated evidence of a positive association between ACE I/D polymorphism with PICH, and suggested that the ACE gene is a PICH susceptible gene in Asian populations. Primary intracerebral hemorrhage (PICH) represents about 15% of all occurring strokes, and has much higher mortality rate than ischemic stroke. In the United States approximately 20,000–40,000 new cases of PICH emerge each year Growing evidence implicates angiotensin–converting enzyme (ACE), a key enzyme of the rennin-angiotensin system, as an important modulator of cerebrovascular disease (CVD). The ACE gene is located on chromosome 17q23 and consists of 26 exons and 25 introns. A crucial polymorphism termed Ins/Del occurs due to an insertion (I) or deletion (D) of 287 base pairs within intron 16. The D allele of the polymorphism accounts for significantly higher serum ACE levels and activity, compared to carriers of the insertion allele In view of few modifiable risk factors for PICH, acute therapeutic options are limited to controlling hypertension. Improving risk assessments through the identification of genetic variants in ICH may contribute to prevention efforts. A previous systematic review of hemorrhagic stroke, published in 2008, failed to confirm a significant and consistent association with ACE I/D Here, we conducted a comprehensive literature-based meta-analysis in which we updated the sample size and explored the ACE I/D variants association with PICH.We independently searched papers using Medline (PubMed) and Embase through Dec 2012 to identify eligible genetic association studies evaluating ACE genotype and PICH. Search terms included hemorrhagic stroke, intracerebral hemorrhage and cerebral amyloid angiopathy, in combination with polymorphism or SNP. Search results were limited to articles of human studies written in English. All references cited in the identified publications were also searched for additional studies not indexed in these two databases. Case reports, editorials, and review articles were excluded.The inclusion criteria were: (1) case–control studies and cohorts which evaluated the association between ACE I/D polymorphism and PICH risk; (2) studies using validated molecular methods for genotyping; and (3) studies containing independent data of ACE I/D genotype frequency for estimating an odds ratio (OR) with a 95% confidence interval (CI). Cases were confirmed with diagnosis of intracerebral hemorrhage based on neuroimaging (magnetic resonance imaging or computerized tomography). Control populations: The vast majority of subjects classified as ‘controls’ were not subjected to neuroimaging, based on the lack of symptoms present following a clinical assessment for hemorrhage.Studies were excluded if (1) the patients were children (age<18 years), (2) the genotype frequency was not adequately reported (and such data could not be obtained from the authors), (3) a case was caused by trauma, hemorrhagic rupture of a tumor or arteriovenous malformation, (4) the study focused solely on intracranial hemorrhage following acute ischemic stroke.For studies containing overlapping cases or controls, the most recent and/or largest study with extractable data was included to avoid double counting.From each selected study the following information was extracted: name of first author, journal, year of publication, country, ethnicity, demographic characteristics, number of ICH cases and controls, confirmation of diagnosis, genotyping method, genotype distributions and allele frequency, and whether genotypes were stated to be in Hardy-Weinberg equilibrium (HWE). If the frequencies of the alleles and/or the genotypic distributions were unavailable, detail information was calculated from the papers. We did not contact individual authors for further information.The initial data extraction was undertaken by Y Liu. Several subsequent passes were then confirmed by Y Sun to ensure the comprehensive inclusion of all appropriate studies. All studies were adjudicated by 2 independent reviewers (Q Sun and Z Zhong), with disagreements resolved through discussion and consensus.Two investigators (Y Liu and Y Sun) independently rated the quality of each retrieved study using a 10-point scoring system developed for this meta-analysis. The scoring system was based on factors that we believed would be indicators of high quality observational studies . Study d2 based Q statistic and I2 with statistical significant considered at P<0.10 A list of details extracted from each study is included in The Begg’s funnel plots and Egger’s test were conducted to statistically assess potential publication bias  A flow diagram of the performed database search is shown in The included studies were published from 1996 through 2011. All 8 studies evaluated susceptibility to PICH. The characteristics of each study are summarized in Genotype and allele distributions for each study are shown in In addition, the Z value for the test for overall effect was 2.28 (P = 0.022) for DD vs. DI+II model, 1.22 (P = 0.224) for DD+DI vs. II model, 1.75 (P = 0.080) for D allele vs. I allele, suggesting that subjects who carry the DD genotype are more vulnerable to PICH compared to those in the dominant model. Overall, the results suggest that the variant genotypes of ACE I/D polymorphism were associated with a significantly higher risk of PICH when all the eligible studies were pooled into the meta-analysis.dominant = 0.018, Precessive = 0.070, Pallele = 0.006). The pooled OR from the two studies did not show a significant association for any genetic model . Although ethnic background produced a significant association, the ethnic effect in the meta regression was not significant (P = 0.483). For the gender difference, we found more males than females were included in the meta-analysis for both PICH patients and controls. There was no significant effect of gender in the meta regression analysis (P = 0.958).To assess the stability of the results of the current meta-analysis, we performed sensitivity analysis by sequentially excluding each study. The pooled ORs ranged from 1.40 to 1.82 in the dominant model, and statistically similar results were obtained after sequentially excluding each study except for the Slowik’s or Li’s study. The pooled ORs ranged from 1.14–1.42 in the recessive model, and statistically similar results were obtained after sequentially excluding each study except for the Catto study. The pooled ORs ranged from 1.18–1.41 in the allele contrast model, and statistically similar results were obtained after sequentially excluding each study except for the Catto or Nakata studies. These results suggested that the stability of our meta-analysis is affected by these four studies.Publication bias was assessed by the Begg’s funnel plot and Egger’s test. Visual inspection of the shape of the funnel plots showed symmetry in the DD vs. II+DI comparison genetic model, suggesting the absence of publication bias . SubsequThe renin angiotensin system has been implicated in the pathophysiology of PICH by accelerating atherosclerosis The use of composite quality scales may produce inherent problems and make it difficult to interpret the results, however the scale seemed to provide an overall assessment when comparing studies Considering that the role of ACE in the pathogenesis of cerebrovascular lesions remains unclear, our meta-analysis findings may account for the relationship of ACE polymorphism with PICH. The ACE D allele has been described to be associated with 28–47% of the variance of ACE activity both in the circulation Subgroup analysis revealed a significant increased risk of PICH among Asians, but not in Caucasians, suggesting that ethnic differences in genetic backgrounds and the environment they lived in may be involve in the incidence of PICH. There are a number of factors that might explain the differences seen in the two populations studied. First, genetic backgrounds (SNP frequency) vary greatly from Asians to Caucasians. Second, the difference in living environments between these two populations may account for the different genetic effects. Thus, additional studies using different population are warranted to further validate ethnic differences on the impact of this polymorphism on PICH risk.Only one study on women identified a positive association, therefore there was not enough power to investigate a potential gender interaction for the ACE I/D variant. Such a sex-specific influence may attribute to corticosteroid hormones, which play a crucial role in the renin-angiotensin-aldosterone system in a variety of tissues A potential publication bias may be present in the allele genetic model. Nevertheless, the combined risk estimate may not be changed just by modifying this bias using the trim and fill method. Though no significant publication bias for DD vs DI/II was found using the Begg’s funnel plots or the Egger’s test, the results should be explained with caution. Interpreting the p value in the setting of significant heterogeneity between studies may be problematic and limited.There is a known positive association between hypertension and ICH Some limitations have already been considered when explaining the results in this meta-analysis. First, our study is under-powered to some degree due to its small sample size In conclusion, the present study provides the most comprehensive meta-analysis to date on the assessment of the relationship between the ACE I/D polymorphism and PICH risk. The present study supports an association between ACE I/D polymorphic variant and PICH in Asians, but not in Caucasians. However, additional large-scale studies with well-matched controls and multi-ethnic populations are required to validate our conclusions.Figure S1Forest plot of OR with 95% CI for ACE I/D polymorphism in PICH susceptibility. (recessive model).(TIF)Click here for additional data file.Figure S2Forest plot of OR with 95% CI for ACE I/D polymorphism in PICH susceptibility. .(TIF)Click here for additional data file.Figure S3Begg’s funnel plot for publication bias in selection of studies of the ACE I/D polymorphism. (recessive model).(TIF)Click here for additional data file.Figure S4Begg’s funnel plot for publication bias in selection of studies of the ACE I/D polymorphism. .(TIF)Click here for additional data file.


Previous investigations of glioma risk in women have focused on oral contraceptive (OC), hormone replacement therapy (HRT), and reproductive factors. However, the results of published studies were inconclusive and inconsistent. Thus, a meta-analysis based on published case-control studies was performed to assess the role of exogenous and endogenous hormones factors in glioma risk.The PubMed and EMBASE databases were searched without any restrictions on language or publication year. Reference lists from retrieved articles were also reviewed. We included case-control studies reporting relative risks (RRs) with corresponding 95% confidence intervals (CIs)  between oral contraceptive (OC) and hormone replacement therapy (HRT) use, reproductive factors and glioma. Random-effects models were used to calculate the summary risk estimates.Finally, 11 eligible studies with 4860 cases and 14,740 controls were identified. A lower risk of glioma was observed among women who were ever users of exogenous hormones  compared with never users. An increased glioma risk was associated with older age at menarche . No association was observed for menopause status, parous status, age at menopause, or age at first birth and glioma risk.The results of our study support the hypothesis female sex hormones play a role in the development of glioma in women. Additional studies are warranted to validate the conclusion from this meta-analysis and clarity the underlying mechanisms. Glioma is the most common type of central nervous system tumor, which accounts for more than 70% of cases We searched the PubMed and EMBASE databases, using the terms “(glioma OR brain cancer OR brain neoplasms OR brain tumor) AND  OR hormone replacement therapy (HRT) OR menopausal hormone therapy OR climacteric OR reproductive history) AND (risk assessment OR risk OR risk factors)”. The citations of the identified articles were also screened for additional studies. We neither imposed any limitations on language or publication year nor sent e-mails to the corresponding authors of included studies to retrieve the original data. The latest search was performed on October 1, 2012.Studies were considered eligible if the studies met the following inclusion criteria: (1) a case-control study; (2) tested the association between OC, HRT, reproductive factors and glioma risk; (3) provided estimates of relative risk with corresponding 95% CIs ; (4) in case of multiple reports of the same trial, we selected the most recent publication with the largest number of subjects.Two authors carefully and independently extracted the following data from each available study: the first author’s last name, publication year, country in which conducted, study period, age of subjects, study design, size of the study population (case/control), proxy interview (yes/no), data collection, exposure variables and categories. Any disagreements were resolved by discussion.The Newcastle-Ottawa Scale (NOS), a validated modality for evaluating observational and non-randomized studies, was used to assess the quality of included studies. This measure assesses aspects of methodology in observational studies related to study quality, including selection, comparability and exposure or outcome. The NOS ranges from zero to nine stars and the studies with ≥6 * were considered to be of relatively higher quality.The STATA software, version 11.0  was used to compute pooled RRs and 95% CIs, generate forest plots, determine whether there was a statistical association, evaluate heterogeneity, perform a sensitivity analysis, and investigate publication bias 2 test was also used. The I2 statistic yields results ranged from 0 to 100%  The RR was used as the measure of association across studies. Because glioma is a rare disease, ORs (odds ratios) were deemed equivalent to RRs. Therefore, we reported all results as relative risk for simplicity. The most-adjusted risk estimates were retrieved and used in the meta-analyses; however, when unavailable, unadjusted risk estimates were used. The unadjusted risk estimates were extracted directly from the article or computed from the exposure distributions for cases and controls given in the papers. The relative risks and corresponding standard errors (derived from the CIs) from individual studies were transformed to their natural logarithms to stabilize the variances and to normalize the distributions. Heterogeneity between studies was evaluated by Cochran’s Q statistic and heterogeneity was considered significant when P<0.1 Sensitivity analysis and publication bias analysis were performed to assess the stability of the results as previously For all exposure variables and categories, we adopted criteria from original publications. In this meta-analysis, we compared ever users with never users for OC and HRT. For HRT, one study reported the risk estimates of using hormones to treat gynecologic problems To assess the effect of reproductive factors, we performed a meta-analysis of the comparison of the highest versus lowest category in each study. For age at menarche and menopause, three studies reported that the oldest age at menopause was compared to the second tertile 2 = 0.0%). In Hochberg and colleagues’ research, it is not entirely clear that all subjects in this study are female. Therefore, we excluded this study and the cumulative risk estimate was 0.700 (95% CI: 0.596–0.822) (Risk estimates for OC ever versus never use were reported in six studies 6–0.822) . The poo2 = 0.0%).For HRT, six studies were included in this meta-analysis 2 = 62.4%).Associations of glioma risk with menopausal status were reported in five studies 2 = 69.9%).Six studies analyzed the role of pregnancy status on glioma risk 2 = 57.6%).Six studies examined the association of glioma and age at menarche 2 = 0.0%).Five articles provided information on glioma risk with age at menopause 2 = 45.1%).The effect of age at first birth was examined by six studies All results of the meta-analysis were also presented in In the sensitivity analysis, we excluded one single study at a time to investigate the influence of each single study on the overall risk estimate. For parous status, the combined RR for parous versus nulliparous was 0.797 (95% CI: 0.637–0.997)  after exThe results of Egger’s test suggested there was no evidence of publication bias  were not robust. Third, measurement error associated with exposure should be noted. Almost all studies included in our meta-analysis used a certain percentage of proxy-reporting measures. Hence, the pooled estimation provided less clear results. Fourth, potential publication bias might influence the findings, yet little evidence of publication bias was observed in the present meta-analysis. Finally, because the results of the current meta-analysis mainly involved Western populations, additional research in other populations is warranted to generalize the findings.In conclusion, a decreased risk of glioma with OC or HRT use was observed, whereas the risk of glioma increased with older age at menarche. Therefore, we supposed that female sex hormones closely relate with the risk of glioma. Further prospective research is warranted to extend this finding by increasing the sample sizes and considering multiple factors.Figure S1Forest plot of OC use (ever vs. never) and glioma risk after excluding Hochberg and colleagues’ research.(TIF)Click here for additional data file.Table S1Methodological quality of included studies based on the Newcastle–Ottawa Scale for assessing the quality of included studies.(DOCX)Click here for additional data file.


Many case-control studies have been performed in the recent past to investigate the association between CCL5 -403 G>A (rs2107538) gene polymorphism and tuberculosis (TB) susceptibility in various ethnic groups. However, these studies have produced inconsistent and contradictory results. In the present study, meta-analysis was performed to assess the association between CCL5 -403 G>A polymorphism and TB risk.Quantitative synthesis was done for the published studies based upon association between CCL5 -403 G>A polymorphism and TB risk from PubMed (Medline), EMBASE web search. Pooled odds ratios (ORs) and 95% confidence intervals (95% CIs) were calculated for allele contrast, homozygous, heterozygous, dominant and recessive genetic models.A total of six studies comprising 1638 confirmed TB cases and 1519 healthy controls were included in this meta-analysis. Variant A allele  and variant homozygous  carriers were significantly associated with TB susceptibility. Similarly, recessive model  also indicated increased TB risk. Whereas, heterozygous  and dominant  models failed to show increased risk of developing TB.This meta-analysis suggests that there is a significant association between the CCL5 -403 G>A polymorphism and increased risk of TB. However, larger well-designed epidemiological studies with stratified case control and biological characterization may be helpful to validate this association. Mycobacterium tuberculosis (M. tuberculosis) and it leads to approximately 1.5 million annual deaths globally M. tuberculosis infection, only 5–15% of people develop active TB disease during their lifetime M. tuberculosis has the ability to survive within the host phagocytic cells, and the relation between the host and the bacteria may lead to tissue damage characterized by granuloma formation, tissue necrosis along with development of cavities and, lastly, spreading of the disease Tuberculosis (TB) is one of the commonest infectious disease, and remains a major public health concern owing to spread epidemically in many parts of the world Chemotactic chemokine (C-C motif) ligand 5 (CCL5) belongs to the family of CC chemokines, and considered as a major chemokine, mostly involved in immunoregulatory and inflammatory activities owing to their ability to recruit, activate and co-stimulate T-cells and monocytes Many functional polymorphisms in the CCL5 gene have been deciphered in past, among them -403 resulting in a G>A (rs2107538) polymorphism have been identified in the promoter region and shown association with altered transcriptional activity and subsequent expression in human cell line A systematic search was carried out through PubMed (Medline), EMBASE web data-bases covering all research articles published with a combination of the following key words: “CCL5 OR RANTES gene (polymorphism OR mutation OR variant) AND tuberculosis or TB (last updated on March 2013). Potentially relevant genetic association studies were evaluated by examining their titles and abstracts, and all published studies matching the selected eligible criteria were retrieved and incorporated in this study.In order to minimize heterogeneity and to facilitate the proper interpretation of our findings, studies included in the present meta-analysis had to meet all the following criteria: a) assessed the association between -403 G>A polymorphism and susceptibility of TB, b) used a case-control design based on unrelated individuals, c) recruited pathologically confirmed TB patients and TB free controls, d) have available genotype frequency in cases and controls, e) and published in the English language. In addition, when the case-control study was included by more than one article using the same case series, selection was done for the study that included the largest number of individuals. The main reasons for exclusion of studies were, a) overlapping data and b) case-only studies, c) review articles, d) genotype frequencies or numbers not reported. The supporting flowchart  of studiFor each research publication, the methodological quality assessment and data extraction were independently abstracted in duplicate by two independent investigators using a standard protocol and data-collection form according to the inclusion criteria listed above to ensure the accuracy of the data. In case of disagreement on any item of the retrieved data, the problem was fully discussed to reach an agreement. Characteristics abstracted from the studies included first author’s name, year of publication of the report, the country of origin, source and number of the cases and the controls, genotype frequencies, and type of the study.2 statistics was utilized to quantify inter-study variability, and the larger values showed an increasing degree of heterogeneity http://www.meta-analysis.com/pages/comparisons.html for the selection and utilization of CMA V2.Pooled odd ratios (ORs) and their corresponding 95% class intervals (CIs) were calculated to examine the relation between CCL5 -403 G>A polymorphism and TB risk. Heterogeneity assumption was measured by the chi-square based Q-test A total of twenty four articles were retrieved by literature search from the PubMed (Medline), EMBASE web-based data-bases. All retrieved articles were assessed by reading their titles and abstracts. The full texts for the potentially relevant articles were further evaluated for their inclusion and suitability for this meta-analysis. In addition to the data-base search, the reference lists of the retrieved articles were further screened for other possible potential studies. As mentioned in the materials and methods section, inclusion and exclusion criteria were set and implemented for the selection of pertinent studies. Studies either using CCL5 polymorphism to predict survival in TB patients or considering CCL5 variants as an indicator for response to therapy were excluded. Research studies measuring the levels of CCL5 mRNA or protein expression or review articles were also excluded. Based upon the selection criteria, only case-control or cohort design based studies having frequency of all three genotypes were included in the current meta-analysis. After cautious screening and following the inclusion and exclusion criteria, six eligible original published studies were included in the present meta-analysis . A detaiIn order to examine the publication bias among the included studies for the meta-analysis, Begg’s funnel plot and Egger’s test were performed. The shape of funnel plots was appearing symmetrical and the results of Egger’s test provided the statistical evidence of the funnel plot. The results demonstrated lack of publication bias among all comparison models .2 statistics were employed to test the heterogeneity among the selected publications. Heterogeneity was observed in all models, such as allele (A vs. G), homozygous (AA vs. GG), heterozygous (GA vs. GG), dominant (AA+GA vs. GG) and recessive (AA vs. GG+GA) genotype model, which were included for the meta-analysis. Thus, the random effect model was applied to calculate the pooled ORs and 95% CI  was employed to evaluate the overall association between the CCL5 -409 G>A polymorphism and TB susceptibility. The pooled results demonstrated that variant allele A was significantly associated with increased risk of TB  . Mutant Despite recent advancements in TB diagnosis and antimycobacterial therapy, the prognosis of TB patients remains miserable. It is well established that risk of TB is determined not only by the infection causing agent and environmental factors but also by the host genetic factors To the best our knowledge, this is the first meta-analysis study addressing the association between CCL5 -403 G>A gene polymorphism and susceptibility of TB, and in this study, a total of six studies were included for the analysis. The pooled results demonstrated that CCL5 -403 G>A polymorphism have substantial effect on the occurrence of TB. Subjects with variant (A) allele and variant homozygous (AA) showed 1.3 and 1.5 fold increased risk of developing TB in comparison with the wild type G allele and homozygous AA genotype. Similarly, recessive model has shown increased risk of TB. CCL5 plays a key role in the antimycobacterial immune response by recruiting mononuclear cells to the site of infection M. tuberculosis infection seems to be polygenic Lastly, the genetic control of the immune response against We conclude that meta-analysis is an extremely valuable and economical method which pools both statistically significant and non-significant results from individual similar studies and produces an absolute conclusion Figure S1PRISMA 2009 Flow Diagram.(TIF)Click here for additional data file.Checklist S1PRISMA 2009 Checklist.(DOC)Click here for additional data file.


Epidemiological studies have shown that tooth loss is associated with risk of head and neck cancer (HNC); however, the results were inconsistent. Therefore, we conducted a meta-analysis to ascertain the relationship between tooth loss and HNC.We searched for relevant observational studies that tested the association between tooth loss and risk of HNC from PubMed and were conducted up to January 30, 2013. Data from the eligible studies were independently extracted by two authors. The meta-analysis was performed using the Comprehensive Meta-Analysis 2.2 software. Sensitivity and subgroup analyses were conducted to evaluate the influence of various inclusions. Publication bias was also detected.Ten articles involving one cohort and ten case-control studies were yielded. Based on random-effects meta-analysis, an association between tooth loss and HNC risk was identified . The sensitivity analysis shows that the result was robust, and publication bias was not detected.Based on the current evidence, tooth loss is probably a significant and dependent risk factor of HNC, which may have a dose-response effect. People who lost six or more teeth should pay attention to symptoms of HNC, and losing 11 teeth or 15 teeth may be the threshold. Head and neck cancer (HNC) mainly originates in the oral cavity, pharynx, and larynx. HNC accounts for 12% of all malignancies worldwide. An estimated total of 400000 cases of oral cavity and pharynx diseases, 160000 cases of laryngeal cancer, and 300000 mortality per year Both HNC and tooth loss share common risk factors; moreover, given their special anatomic location, an interesting assumption was formed on whether or not an association between tooth loss and HNC existed? Zheng et al (1990) Cohort, case-control, and cross-sectional studies that evaluated the association between tooth loss and HNC and those that meet the following criteria were considered eligible for inclusion: (1) full-text could be obtained; (2) clear diagnostic criteria for HNC and definition of tooth loss were reported; and (3) the adjusted and/or unadjusted hazard ratios (HRs), odds ratios (ORs), or relative risks (RRs), and the associated 95% confidence intervals (CIs), or the numbers of events that can calculate these factors, were reported. If more than one study covered the same population, only the report containing the most comprehensive information on that population was included. Two authors independently evaluated the eligibility of all the retrieved studies, and disagreements were resolved by discussion.The PubMed database was searched up to January 30, 2013  for published studies that tested the association between tooth loss and HNC. The search term  AND (“dentition” OR “tooth loss”) was used. We also reviewed the reference lists of included articles and recent reviews.Two authors independently collected and tabulated the following information of each eligible study: the first author’s surname, year of publication, study design, country of origin, sample size, number of events, age range, assessment of tooth loss and HNC, tumor site and pathologic type of HNC, crude or adjusted point estimates and relevant 95% CIs, and the covariates for the adjusted point estimates.0 is the incidence of the outcome of interest in the non-exposed group. The standard error (SE) of the resulting converted OR was then determined using the following formula: The design of most of the included studies was a case-control study and reported ORs. Only one study was prospective cohort and reported HR + teeth, 15+ teeth, and 20+ teeth.The numbers of lost tooth varied in the included studies; hence, we gathered and categorized these teeth into five categories as follows: lost 1 to 6 teeth, 6 to 15 teeth, 112 statistics 2 values of 25%, 50%, and 75%, respectively 2≤25%, which indicates no evidence of heterogeneity, we used the fixed-effect model; otherwise, we used the random-effects model. In the presence of heterogeneity, we performed sensitivity analysis to explore the possible explanations for heterogeneity by removing each study in each turn to test the robustness of the main results or by switching the fixed and random effects models.We computed a pooled OR and relevant 95% CI by using the Comprehensive Meta-Analysis software, version 2.2  We used the Stata 12.0 software for the dose-response estimates based on pooled ORs and 95% CIs by each category of the number of lost teeth. Publication bias was assessed by visual inspection of the funnel plots and the Egger linear regression test From the 82 records initially found, 10 articles involving 11 case-control studies From the included studies, the study of Guha et al. 2 = 85.59%). Nine articles involving 10 studies 2 = 82.92%). The result from the random-effects model showed that 6 to 15 teeth loss could significantly increase the risk of developing HNC by 1.58 times  of 11+ teeth loss and the risk of HNC. A significantly increased risk for developing HNC by 1.63 times was observed . The pooled result of the three case-control studies + teeth loss could increase the risk of HNC by 1.89 times .Three studies reported 1 to 6 teeth loss and risk of HNC. Among the three studies, one + teeth loss, and from 1.48  to 1.90  of 15+ teeth loss , and the adjustment for confounding factors. Our sensitivity analysis  and subgroup analysis  results provide evidence that country of origin and definition of reference group probably contributed to the heterogeneity.The major strength of our study is that this it is the first to perform meta-analysis on this topic. We searched relevant published studies via electronic and hand searching. To the best of our knowledge, we have collected all published studies that met the inclusion criteria and the publication bias test also provided no evidence of publication bias. Moreover, we performed sensitivity analysis by using three methods and subgroups analysis based on the study design, definition of the reference group, country of origin, which could improve the reliability of the results and reduce the performance bias of the meta-analysis. Third, the association of tooth loss with the risk of HNC persisted and remained without substantial change in the sensitivity analyses based on various methods. In addition, the subgroup analysis results indicate that, with accumulating evidence and enlarged sample sizes, the statistical power is enhanced to provide more precise and reliable risk estimates. Finally, our results were based on adjusted estimates, thereby making the result more credible than unadjusted ones.http://www.ncbi.nlm.nih.gov/pubmed/23990929). Their work is perfect! However, when we compared our meta-analysis to theirs, we found there were three major differences/advantages of our meta-analysis. First, the deadline of search of theirs was March, 2013 and yielded eight case-control studies and one cross-sectional study; however, the deadline of search in our meta-analysis was January 30, 2013 and yielded 10 articles involving 11 case-control studies and one cohort study. Obviously, our search is more comprehensive. Second, we performed meta-analysis based on the number of lost tooth Click here for additional data file.


GSTM1) and glutathione S-transferase theta-1 (GSTT1) null polymorphisms and childhood acute leukemia; however, the conclusions remain controversial. We performed an extensive meta-analysis on 26 published case-control studies with a total of 3252 cases and 5024 controls. Crude odds ratios (ORs) with 95% confidence interval were used to assess the strength of association between childhood acute leukemia risk and polymorphisms of GSTM1 and GSTT1. With respect to GSTM1 polymorphism, significantly increased risk of childhood acute leukemia was observed in the overall analysis . Furthermore, a stratification analysis showed that the risk of GSTM1 polymorphism are associated with childhood acute leukemia in group of Asians , Blacks , ALL , ‘< 100 cases and <100 controls’ , ‘≥ 100 cases and ≥ 100 controls’ , and population-based control source . With respect to GSTT1 polymorphism, significant association with childhood acute leukemia risk was only found in subgroup of Asian. This meta-analysis supports that GSTM1 null polymorphism is capable of causing childhood acute leukemia susceptibility. Several molecular epidemiological studies have been conducted to examine the association between glutathione S-transferase mu-1 ( GSTM1) and glutathione S-transferase theta-1 (GSTT1) genes might have an increased risk of the development of childhood acute leukemia. Leukemia is the most common form of cancer in childhood accounting for approximately one third of all childhood cancers ; which iGSTM1 gene, is found on the chromosome 1p13.3 [GSTT1 gene, is locate on the chromosome 22q11.23 [GSTM1 and GSTT1 have absent activity of the respective enzyme. DNA-adduct formation and rates of somatic mutation have been reported to be increased in carriers of null alleles [GSTM1 and GSTT1 null genotypes has been associated with several types of malignancies, including stomach cancer [The GSTM1 and GSTT1 are phase II metabolic enzymes have the ablity to detoxify a wide variety of electrophilic compounds including the activated carcinogens. Human glutathione S-transferases are divided into eight distinct classes as alpha, kappa, mu, omega, pi, sigma, theta, and zeta based on amino acid sequence similarity and antibody cross-reactivity ,4. The me 1p13.3 . The the22q11.23 . Homozyg alleles . Individ alleles ,9. An inh cancer , lung cah cancer , bladderh cancer , cervicah cancer , and acuh cancer . GSTM1, GSTT1 gene and childhood acute leukemia in diverse populations [GSTM1 and GSTT1 of small effect sizes. To estimate the effect of GSTM1 and GSTT1 polymorphisms on the childhood acute leukemia, as well as to quantify the potential between-study heterogeneity, we conducted a meta-analysis on 26 published case-control studies with a total of 3252 cases and 5024 controls.GST polymorphisms were first reported as risk factors for childhood acute leukemia in 1997 . Since tulations –33. HoweGSTM1 and GSTT1 and the risk of childhood acute leukemia were identified by searching for articles in the PubMed and Chinese Biomedical Literature Database until 1 March 2013. Various combinations of the search terms ‘(GSTM1 or GSTT1) and (polymorphism or polymorphisms) and childhood acute leukemia’ were used to screen for potentially relevant studies. Additional articles were also checked using the references cited in these publications. Articles that had data on the different types of childhood acute leukemia , acute myeloid leukemia (AML) and acute non-lymphoblastic leukemia (ANLL)) or different ethnic groups  were treated as independent studies. Studies included in our meta-analysis had to meet all of the following criteria: (i) studied on human beings; (ii) in a case-control study design; and (iii) had detailed genotype frequency of cases and controls or could be calculated from the article text. In current study, data for meta-analysis were available from 17 articles (26 independent case-control studies), including 3252 cases and 5024 controls. Studies addressing the association between polymorphisms of Two independent researchers extracted raw data according to the inclusion criteria. If the two investigators generated different results, they would check the data again and have a discussion to make an agreement. If they could not reach an agreement, an expert was invited to the discussion. Data extracted from the selected articles included the first author’s name, year of publication, country of origin, ethnicity, subtype of acute leukemia, number of cases and controls, genotype frequency for cases and controls, and source of controls.GSTM1 and GSTT1 genes were estimated for each study by odds ratio (OR), together with its 95% confidence interval (CI), respectively. Most studies evaluated GSTM1 and GSTT1 as presence/absence of gene deletion, so that meta-analysis of these polymorphisms were performed using a crude OR (null vs. present). A fixed-effect model using the Mantel-Haenszel method and a random-effects model using the DerSimonian and Laird method were used to combine values from studies. If the P value for heterogeneity was > 0.10 and I2 < 50%, indicating an absence of heterogeneity between studies, we used the fixed-effect model to evaluate the summary ORs. In contrast, if the P value for heterogeneity was ≤ 0.10 or I2 ≥ 50%, indicating a high extent of heterogeneity between studies, we used the random-effect model to evaluate the summary ORs.The risk of childhood acute leukemia that is associated with the polymorphisms of Subgroup analyses were conducted by ethnicity , subtype of acute leukemia , number of cases and controls  and control source . Possible publication bias was tested by Begg’s funnel plot and Egger’s test. All analyses were performed using STATA software, version 9.2 .There were 117 articles relevant to searching strategy. The flow chart shown in A total of 26 studies were included in the meta-analysis with 3252 cases and 5024 controls. Cases consisted of 87.1% patients with ALL, 11.1% patients with AML and 1.8% patients with ANLL. Most of the controls (60.4%) were population-based participants. Totally, 24 studies met the inclusion criteria and were selected in this meta-analysis with 2934 cases and 4592 controls. Cases consisted of 85.7% patients with ALL, 12.3% patients with AML and 2.0% patients with ANLL. Most of the controls (63.3%) were population-based participants. GSTM1 polymorphism and childhood acute leukemia risk is summarized in GSTM1 polymorphism and childhood acute leukemia was found in all subjects   . When st< 100 cases and <100 controls’  and ‘≥ 100 cases and ≥ 100 controls’ . Additionally, subgroup analysis by source of controls indicated that the null genotype has been associated with an increased risk of childhood acute leukemia in population-based studies  but not in hospital-based studies (Table 3).GSTT1 polymorphism and childhood acute leukemia are listed in GSTT1 polymorphism was associated with a significantly increased risk of childhood acute leukemia in Asians   , while t17-1.93) . Sensitivity analyses were performed to determine whether modification of the inclusion criteria of the meta-analysis affected the final results. Although the sample size for cases and controls in 26 studies with a range from 6 to 616, the corresponding pooled ORs were not qualitatively altered with or without the study of small sample. Similarly, the sensitivity analysis indicated no other single study influenced the pooled ORs materially.GSTM1, the shape of the funnel plot did not reveal any evidence of obvious asymmetry (P = 0.064) (P = 0.016) implied some evidence of publication bias. For GSTT1, both Begg’s test (P = 0.941) and Egger’s test (P = 0.991) did not suggest any evidence of publication bias (Begg’s funnel plot and Egger’s test were performed to assess the publication bias of literatures. = 0.064) . Howeverion bias . GSTM1 and GSTT1 null polymorphisms with childhood acute leukemia risk. We demonstrated that the null polymorphism of GSTM1 was associated with a significant increase in overall childhood acute leukemia risk, whereas the null polymorphism of GSTT1 did not appear to have an overall influence on the susceptibility of childhood acute leukemia. Furthermore, in the stratified analyses of GSTM1 null polymorphism, we found a significant influence on childhood acute leukemia risks in Asian and Black ethnic groups, ALL and population-based controls. However, we failed to find any significant relationships between GSTT1 null polymorphism and childhood acute leukemia risk except in group of Asian. The association of the GSTM1 null polymorphism but not the GSTT1 polymorphism with childhood acute leukemia may be an indication of substrate specificity of GSTM1 in metabolism of agents that are involved in the etiology of childhood acute leukemia.The present meta-analysis, including 3252 cases and 5024 controls from 26 case-control studies, exploring the association of GSTM1 and GSTT1 polymorphisms and childhood acute leukemia risk. The GSTM1 polymorphism is one of the most studied loci relating to childhood acute leukemia risk. The homozygous deletion resulting in functional loss of the GSTM1 enzyme has been implicated in the genesis of several cancers, including cervical neoplasia [GSTM1 null genotype is associated with a higher risk of childhood acute leukemia. In 2005, a meta-analysis by Ye et al. [GSTM1 and GSTT1 with childhood ALL risk, including 4721 subjects . A recent meta-analysis of 15 published case-control studies on the effect of these polymorphisms and risk of childhood ALL was performed [GSTM1 polymorphism but not GSTT1 polymorphism was associated with the risk of childhood ALL [ To the best of our knowledge, we conducted by far the largest and most comprehensive meta-analysis for quantitative analyses between the roles of the eoplasia , coloreceoplasia  and bladeoplasia . The pree et al. , had reperformed . One obshood ALL , is simihood ALL ,36 and nhood ALL . InclusiSeveral factors must be considered in the design of a reliable case-control study in the future. Large sample size with adequate power is one of the most important factors. The choice of the control population is also considered to be a crucial factor because of the possible different exposure to environmental toxicants. Additionally, studies including information on the subtype of childhood acute leukemia are demanded to clarify the relationship between the GST polymorphisms and the subtypes of childhood acute leukemia.There are some limitations should be acknowledged in this meta-analysis. Firstly, in the subgroup analyses of childhood acute leukemia, the number of AML and ANLL subgroups was relatively small, which don't have enough statistical power to explore the real association. Secondly, only three of the examined studies were performed in a Black population, so the ethnicity effect was not adequately investigated. Thirdly, our results were based on unadjusted estimates, while a more precise analysis should be conducted if all individual data was available, which would allow for the adjustment by other co-variants including age, gender, and environmental exposures. Fourthly, childhood acute leukemia is a multi-factorial disease that results from complex interactions between many genetic and environmental factors. It suggests that there will not be single gene or single environmental factor that has large effects on childhood acute leukemia susceptibility. In addition, as in most meta-analyses, publication bias must be considered because only published studies were included in the meta-analysis.GSTM1. It is necessary to conduct large sample studies using standardized unbiased genotyping methods, homogeneous patients with childhood acute leukemia and well matched controls. Additionally, more studies or complete case-control studies, especially stratified by different ethnic background, environmental exposure or other risk factors, should be performed to clarify possible roles of GSTM1 and GSTT1 null polymorphisms in the pathogenesis of childhood acute leukemia in the future.In conclusion, this meta-analysis showed that an increased risk of childhood acute leukemia is associated with the null polymorphism of Figure S1Flow chart of study identification. Studies that had data on the different subtypes of acute leukemia  or different ethnic groups  were treated as independent studies. Thus, a total of 26 studies were included in quantitative synthesis.(TIF)Click here for additional data file.Figure S2Funnel plot analysis to detect publication bias. Each point represents a separate study for the indicated association. Funnel plot for GSTM1 (A) and GSTT1 (B) null polymorphisms in overall analysis.(TIF)Click here for additional data file.Table S1PRISMA Checklist.(DOC)Click here for additional data file.


P = 0.002; TT vs. CC: OR = 0.74, 95 % CI = 0.61–0.91, P = 0.004, Fig. P = 0.014; TT vs. CC/CT: OR = 0.77, 95 % CI = 0.62–0.94, P = 0.010; TT/TC vs. CC: OR = 0.82, 95 % CI = 0.72–0.95, P = 0.006). In the subgroup analysis by ethnicity, there were also significant associations found between VEGF +936C/T polymorphism and breast cancer risk in Asians and Caucasians. In conclusion, the results of our meta-analysis suggest that the VEGF +936C/T polymorphism is significantly associated with breast cancer development and the VEGF 936T allele carriers may be associated with decreased breast cancer risk.The association between vascular endothelial growth factor (VEGF) +936C/T polymorphism and breast cancer risk has been widely reported, but results were inconsistent. In order to derive a more precise estimation of the relationship, a meta-analysis was performed. Eligible articles were identified through search of databases including PubMed, Embase, and Chinese Biomedical Literature Database (CBM). The association between the VEGF +936C/T polymorphism and breast cancer risk was conducted by odds ratios (ORs) and 95 % confidence intervals (95 % CIs). Finally, a total of 13 studies with 6,879 cases and 7,219 controls were included in our meta-analysis. Overall, a significant association was found between VEGF +936C/T polymorphisms and the risk of breast cancer in overall populations under five models (T vs. C: OR = 0.83, 95 % CI = 0.73–0.94,  Breast cancer is the second most common cancer in women after skin cancer, accounting for about one third of all cancers in women . Global 18F-fluorodeoxyglucose, used for detection and staging of breast cancer [Previous studies have suggested that low-penetrance susceptibility genes combined with environmental factors may be important in the development of cancers, including breast cancer , while vt cancer –18.Up to now, a number of studies have reported the association between the VEGF +936C/T polymorphism and breast cancer susceptibility, but the results remain inconsistent –29. In oWe performed an electronic search of the PubMed, Embase, and Chinese Biomedical Literature Database (CBM) to retrieve articles linking VEGF +936C/T gene polymorphism and susceptibility to breast cancer available until August 2013 with keywords “breast cancer,” “breast neoplasm,” “vascular endothelial growth factor,” “VEGF,” “polymorphism,” and “variant,” and there were no limitations to the language of publications. Additional studies were identified by a hand search of the references of original studies, and review articles were also examined to find additional eligible studies.Eligible studies included in our meta-analysis had to meet the following criteria: (a) they should be case–control studies, (b) they should evaluate the VEGF +936C/T polymorphism and breast cancer risk, (c) they should supply the available genotype frequency in cases and controls, and (d) they should have sufficient published data for estimating an odds ratio (OR) with 95 % confidence interval (CI). The exclusion criteria were as follows: (a) not a case–control study, (b) no usable data reported, (c) duplicate data, (d) abstract, comment, review, and editorial. When multiple publications reported on the same or overlapping data, the most recent or largest population was selected.Two investigators independently extracted the data, based on the inclusion criteria mentioned above. If conflicting evaluations were encountered, an agreement was reached following a discussion; if agreement could not be reached, then a third author was consulted to resolve the debate. The following information were extracted: (a) the name of the first author, (b) year of publication, (c) country of origin, (d) ethnicity, (e) genotyping methods, (f) source of the control group, (g) distribution of genotypes in case and control groups. We also evaluated whether the genotype distributions were in the Hardy–Weinberg equilibrium.Q statistic test. The effect of heterogeneity was quantified by using I2 values as well as P values [I2 value <50 % and P > 0.10, this suggests that obvious heterogeneity does not exist, ORs were pooled by a fixed-effects model (the Mantel–Haenszel method) [The odds ratio (OR) and 95 % confidence interval (CI) were used to evaluate the strength of associations between the VEGF +936C/T polymorphism and the risk of breast cancer according to five genetic models: allele contrast (T vs. C), homozygote (TT vs. CC), heterozygote (TC vs. CC), recessive (TT vs. TC/CC), and dominant (TT/TC vs. CC) models. The heterogeneity was tested by a chi-square-based P values . If I2  method) . Otherwi method) .http://ihg2.helmholtz-muenchen.de/cgibin/hw/hwa1.pl); if P > 0.05, this suggests that the controls followed the HWE balance. Sensitivity analysis was performed to evaluate the stability of the results. A single study involved in the meta-analysis was omitted each time to reflect the influence of the individual data set on the pooled ORs [P < 0.05 was considered statistically significant), the sensitivity analysis was also conducted. Possible publication bias was tested by Egger’s test  [P < 0.05 for any test or model was considered to be statistically significant.The Hardy–Weinberg equilibrium (HWE)  of controled ORs . When thon bias)  and visuon bias)  in the mP > 0.05). The main characteristics of the studies included in the present meta-analysis are listed in Table After careful examination according to the inclusion criteria, a total of 11 publications with 13 studies including 6,879 cases and 7,219 controls were included in our meta-analysis . In the P = 0.002; TT vs. CC: OR = 0.74, 95 % CI = 0.61–0.91, P = 0.004, Fig. P = 0.014; TT vs. CC/CT: OR = 0.77, 95 % CI = 0.62–0.94, P = 0.010; TT/TC vs. CC: OR = 0.82, 95 % CI = 0.72–0.95, P = 0.006). In the subgroup analysis by ethnicity, there were also significant associations found between VEGF +936C/T polymorphisms and breast cancer risk in Asians  and Caucasians .The main results of this meta-analysis and the heterogeneity test were shown in Table Sensitivity analyses were conducted to determine whether modification of the inclusion criteria of the meta-analysis affected the final results. A single study involved in the meta-analysis was deleted each time to reflect the influence of the individual data set on the pooled OR, and the corresponding pooled OR was not materially altered (data not shown), indicating that our results were relatively stable and credible.P > 0.05, data not shown).A funnel plot and Egger’s test were performed to assess publication bias. The funnel plot is relatively straightforward in observing whether the publication bias is present, and Egger’s test was used to provide statistical evidence of symmetries of the plots. The shape of the funnel plot did not reveal any evidence of obvious asymmetry Fig. . SimilarAngiogenesis is essential for the growth of microscopic cancers into larger tumors, while VEGF plays a central role in angiogenesis through a variety of mechanisms such as effects on endothelial cell proliferation, survival, and migration , 38. PreThe present meta-analysis, including 6,879 cases and 7,219 controls, explored the association between the VEGF +936C/T polymorphism and breast cancer risk. The results show that the VEGF +936C/T polymorphism is significantly associated with breast cancer development. Actually, it might act as a protective factor for breast cancer. In the subgroup analyses based on ethnicity, a significant association was found in Asian and Caucasian populations. This may be explained by the fact that cancer is a complicated multigenetic disease and different genetic backgrounds may contribute to the discrepancy.Although a comprehensive analysis was conducted to show the association between MTHFR polymorphism and risk of breast cancer, there are still some limitations that should be pointed out. First, the number of studies and the number of samples included in the meta-analysis were relatively small. Second, in the subgroup analyses, the number of Asians was relatively small, not having enough statistical power to explore the real association. Additionally, no data were available about Africans. Third, the controls were not uniformly defined. Some studies used controls that were population based, while others used hospital-based controls, which may not be representative of the general population. Finally, our results were based on unadjusted estimates, while a more precise analysis should be conducted if individual data were available, which would allow for the adjustment of other co-variants including age, menopausal status, environmental factors, and lifestyle.In conclusion, the results of our meta-analysis suggest that the VEGF +936C/T polymorphism is significantly associated with breast cancer development and the VEGF 936T allele carriers may be associated with decreased breast cancer risk.


Toll-like receptor 4 plays a role in pathogen recognition, and common polymorphisms may alter host susceptibility to infectious diseases. TLR4 896A>G and TLR4 1196C>T) with infectious diseases.To review the association of two common polymorphisms  was used to quantify the impact of TLR4 variants on disease susceptibility. A meta-analysis was undertaken for outcomes reported in >1 study. Eleven of 37 distinct outcomes were significant. TLR4 896 A>G increased risk for all parasitic infections , malaria , brucellosis , cutaneous leishmaniasis , neurocysticercosis , Streptococcus pyogenes tonsillar disease  , typhoid fever  and adult urinary tract infections , but was protective for leprosy . TLR4 1196 C>T effects were similar to TLR4 896 A>G for brucellosis, cutaneous leishmaniasis, leprosy, typhoid fever and S. pyogenes tonsillar disease, and was protective for bacterial vaginosis in pregnancy  and Haemophilus influenzae tonsillar disease . The majority of significant associations were among predominantly Asian populations and significant associations were rare among European populations.The generalized odds ratio metric (ORDepending on the type of infection and population, TLR4 polymorphisms are associated with increased, decreased or no difference in infectious disease. This may be due to differential functional expression of TLR4, the co-segregation of TLR4 variants or a favorable inflammatory response. Mycobacterium tuberculosis [G 1.09 (95% CI 0.13-9.09)] [G 1.50 (95% CI 0.88-2.56)] ,46,47,523-9.09)] ,46 and a8-2.56)] ,55,56. TG 10.14; 95% CI 1.90-54.16) [G 3.10; 95% CI 1.45-6.67) [S. pyogenes tonsillar disease in Greece  [G 2.26; 95% CI 1.01-5.07) [G 0.55;95% CI 0.31-0.98) in the United States  [G 0.23; 95% CI 0.10-0.55) [Haemophilus influenzae tonsillar disease in a Greek pediatric population  [S. pneumoniae (protection) were not confirmed in this analysis with the use of the ORG metric. For outcomes with a single available study, a significant risk was present for cutaneous leishmaniasis in Iran (OR0-54.16) , neurocy45-6.67) , S. pyog36-7.13)  and typh01-5.07) . A signie cases) , leprosy10-0.55) , and Hae17-1.00) . The sig17-1.00) , a risk TLR4 896 A>G and 1196 C>T were examined : 7.22 and 10.14 for cutaneous leishmaniasis, 4.39 and 3.13 for neurocysticercosis, 2.93 and 3.12 for S. pyogenes tonsillar disease, 2.51 and 2.26 for typhoid fever, 0.36 and 0.23 for leprosy. An exception to the rule was H. influenzae tonsillar disease, where the protective effect of TLR4 896 A>G did not reach statistical significance , while 1196 C>T showed significant association .The significant effects were unidirectional and similar in magnitude when both examined , that isTLR4 896 A>G, TLR4 1196 C>T) with infectious diseases. An increased risk was documented for all parasitic infections combined, malaria [S. pyogenes tonsillitis among Greek children [G and 1196 C>T variants when both associations were studied. We performed a systematic literature review to address the potential association of 2 common TLR4 single nucleotide polymorphisms  metric to quantify the magnitude of associations. This metric provides a straightforward interpretation of the relative risk effect, based solely on genotype distribution [G obviates the need for selecting, estimating and interpreting individual genotype contrasts  and their effect. ORG can also be used in meta-analysis of GAS to summarize effects and produce robust results, avoiding the shortcomings of multiple model testing, namely the lack of biologic justification and non-independency of effects [TLR4 896 A>G association with malaria, the combined ORG showed that the probability of having malaria might be 31% higher for subjects having higher mutational load relative to those with lower mutational load . The application of the ORG metric also resulted in a more conservative estimate of associations, given that associations for infections such as osteomyelitis (39) and S. pneumonia (43) were downgraded to insignificant. The associations derived from tuberculosis data were insignificant similar to those reported [In this study we utilized the generalized odds ratio  among African populations, with only 2% having TLR4 1196 C>T co-segregation. Two studies (on typhoid fever and leprosy) indicated weak linkage disequilibrium in Malaysian [Among Indo-European populations, 6-14% of the individuals are double heterozygous for both polymorphisms . It is salaysian  and Eastalaysian . These dPorphyromonas gingivalis, Fusobacterium nucleatum and Aggregatibacter actinomycetemcomitans [G and 1196 C>T in Europeans [Our analysis on the impact of these polymorphisms in periodontitis illustrates the different impact of polymorphisms based on the population. More specifically, despite the bulk of studies on aggressive and chronic periodontitis, TLR4 variants did not show any significant association, even though TLR4 has been shown to be overexpressed in gingival epithelial cells and gingival fibroblasts -97 in ascomitans -101. Oneuropeans . Importantly, our analysis highlights the need to evaluate the impact of these polymorphisms in different populations and various clinical conditions. Moreover, the absence of significant associations in meta-analysis data for periodontitis, tuberculosis, meningococcal disease and sepsis, signifies that the functional alterations related to polymorphic TLR4 variants may not be critical to produce the clinical phenotype. Lack of reproducibility stands as a barrier for conclusive evidence, and design, sample size and environmental and genetic heterogeneity between populations may affect results. Finally, the presence of a significant effect may rely on the magnitude of functional expression of TLR4. Protection or risk may be moderated by the level of TLR4 functional expression, which is modulated by TLR4 polymorphism and MD-2 presence ,15. TherThe heterogeneity of the populations studied along with multiple endpoints should also be considered as potential study limitation that may influence statistical power. Moreover, different populations mount diverse immunologic responses and the clinical relevance of polymorphisms is not always straightforward. The lack of association for a disease phenotype highlights that gene-to-gene interactions and gene-environment interactions may be influential parameters of disease association. Case-control design of individual GAS precludes adjusted analysis for gene-gene-environment interactions and may have reduced the efficiency of genetic risk estimates, though it is unlikely to inflate false-positive results .Despite these limitations, genetic markers of immune response such as TLR4 variants, are valuable not only to classify high-risk patients based on disease susceptibility but also to predict disease severity and other sequelae. The associations of TLR4 896A>G with hearing loss in survivors of bacterial meningitis  and the  In conclusion, our analysis highlights the complex effect of TLR variants in susceptibility to infectious disease. Some of the effects, such as in malaria, are validated in a variety of studies, whereas single case-control studies should be cautiously interpreted until more information on the specific outcomes is added. Taken in their totality, our results indicate that depending on the infection and the population studied, the same polymorphism may be associated with risk, protection or have no effect. In this context, our analysis provides the rationale for understanding the protective or adverse effect of TLR4 polymorphisms and may provide a basis to explain the maintenance of these polymorphisms.Checklist S1PRISMA checklist.(DOC)Click here for additional data file.Table S1Flow diagram of meta-analysis.(DOCX)Click here for additional data file.


Use systematic review methods to quantify the association between prostatitis and prostate cancer, under both fixed and random effects model. Case control studies of prostate cancer with information on prostatitis history. All studies published between 1990-2012, were collected to calculate a pooled odds ratio. Selection criteria: the selection criteria are as follows: human case control studies; published from May 1990 to July 2012; containing number of prostatitis, and prostate cancer cases. In total, 20 case control studies were included. A significant association between prostatitis and prostate cancer was found, under both fixed effect model , and random effects model . Personal interview based case control studies showed a high level of association , compared with clinical based studies . Additionally, pooled ORs, were calculated for each decade. In a fixed effect model: 1990’s: OR=1.58, 95% CI: 1.35-1.84; 2000’s: OR=1.59, 95% CI: 1.40-1.79; 2010’s: OR=1.37, 95% CI: 1.22-1.56. In a random effects model: 1990’s: OR=1.98, 95% CI: 1.08-3.62; 2000’s: OR=1.64, 95% CI: 1.23-2.19; 2010’s: OR=1.34, 95% CI: 1.03-1.73. Finally a meta-analysis stratified by each country was conducted. In fixed effect models, U.S: pooled OR =1.45, 95%CI: 1.34-1.57; China: pooled OR =4.67, 95%CI: 3.08-7.07; Cuba: pooled OR =1.43, 95%CI: 1.00-2.04; Italy: pooled OR =0.61, 95%CI: 0.13-2.90. In random effects model, U.S: pooled OR=1.50, 95%CI: 1.25-1.80; China: pooled OR =4.67, 95%CI: 3.08-7.07; Cuba: pooled OR =1.43, 95%CI: 1.00-2.04; Italy: pooled OR =0.61, 95%CI: 0.13-2.90.CONCLUSIONS: the present meta-analysis provides the statistical evidence that the association between prostatitis and prostate cancer is significant. Today, inflammation is present in approximately 17% of all cancer cases; findingProstatitis, is defined as inflammation of the prostate gland. According to the prostatitis diagnosis guideline, prostatitis could be classified as acute bacterial prostatitis, chronic bacterial prostatitis, inflammatory prostatitis , noninflammatory prostatitis and asymptomatic prostatitis. ProstatMeta analysis is a quantitative systematic method to test the effectiveness of exposure/treatment, in both cohort and case control studies. A previous meta analysis involvinWe conducted a literature search for ‘prostatitis’ and ‘prostate cancer’, or ‘inflammation’ and ‘prostate cancer’, or ‘meta’ and ‘prostatitis’ and ‘prostate cancer’ in both PubMed and Medline databases, based on English literature. In addition, the references and citations of studies were also reviewed. If cited or referenced articles were eligible, they would be included into pooled studies candidates in order to prevent the loss of any important and useful data. The selection criteria were as follows: human case control studies; published from May 1990 to July 2012; that contained number of prostatitis and prostate cancer cases. The study also must provide comprehensive information including: age, race/country, number of people with/without prostatitis, number of people with/without prostate cancer, study period, and data source . Additionally, in all qualified studies, the patients who were diagnosed and managed as prostatitis must meet the criteria for the NIH prostatitis guideline,20–39. Quantitative Meta-analysis was performed using Stata 11. Data were imported under study name expressed by the last name of authors, year of study, country, race, data source, number of cases (cancer), number of controls, number of cases exposure (prostatitis) and number of unexposed cases not under exposure. One concern was that prostatitis has several classifications, including chronic bacterial prostatitis, acute prostatitis, and asymptomatic prostatitis. However, in our 20 studies, numbers of general prostatitis are given instead of numbers of different disease classification, and we are unable report related statistical results based on classification.Odds ratios were calculated using the given information. Pooled log OR were calculated under a fixed effect model, which assume exposure’s effect on cancer cases is same across studies. Heterogeneity of ORs was tested. If p-value of heterogeneity was significant (<0.05), random effects model was more appropriate, which assumes that exposure’s effects on cancer cases are different across studies. Statistical methods in Meta-analysis are desired in the following literature review. In our study, we use both fixed effect and random effects model and provide related results in tables and figures.Stata version 11  was the main statistical tool used in this study. Both fixed effect and random effects model were used to test the effect of prostatitis in increasing the risk of prostate cancer. The fixed effect model assumes that the effect of treatment is the same among all studies, and the random effects model assumes that the effects might be different, and therefore, random effects model usually has a wider confidence interval. When heterogeneity is significant, it indicates that a random effects model is more appropriate than a fixed effect model.Data were stratified based on the data source , country, and decades . All the stratification is tested to identify the bias and difference between different groups, based on the assumption of heterogeneity (fixed effect or random effects group). Publication bias was tested based on Egger’s test, to exam whether the bias is between the published and unpublished papers. Funnel plots were also graphed to check the small study effect. In This study demonstrates that there is a significant positive relationship between prostatitis and prostate cancer, under both fixed effect and random effects model. This conclusion agrees with many previous biological and epidemiological studies mentioned above ,6–10, whFrom Publication bias and small study effect were also considered in this study. According to Egger’s publication bias plot, there was no evidence that unpublished papers and studies have significant effects on this study (p=0.820), with the intercept=0.912. Similar results were obtained when using Begg’s Funnel plot, where small studies are widely spread at the bottom of funnel. Thus, funnel plots are almost symmetric, and no small study effects observed. There are some limitations of this study. First, due to restriction placed on language and datasets, there are more studies, which were not included in this study, causing a potential bias. In addition, unpublished papers, especially ones without significant results, may still have effects on the overall pooled OR. Second, prostatitis is a clinically progressive (acute prostatitis can become chronic prostatitis), multi-classification disease, according to the prostatitis diagnosis guideline. Most of the studies we considered did not divide the prostatitis cases into different classification, and that may be a problem if all types of prostatitis are related to prostate cancer. Prostate cancer is also associated with age and race. However due to limited numbers of cases and controls within each age and race stratum, this study failed to find a relationship between age, race. Checklist S1Prisma checklist.(DOC)Click here for additional data file.