Here we conducted a meta-analysis to verify the effect of rs3775291 on AMD. We searched for genetic association studies published in PubMed, EMBASE and Web of Science from start dates to March 10, 2015. Totally 235 reports were retrieved and 9 studies were included for meta-analysis, involving 7400 cases and 13579 controls. Summary odds ratios (ORs) with 95% confidence intervals (CIs) for alleles and genotypes were estimated. TLR3 rs3775291 was associated with both geographic atrophy (GA) and neovascular AMD (nAMD), with marginally significant pooled-P values. Stratification analysis by ethnicity indicated that rs3775291 was associated with all forms of AMD, GA and nAMD only in Caucasians  but not in East Asians. However, the associations could not withstand Bonferroni correction. This meta-analysis has thus revealed suggestive evidence for TLR3 rs3775291 as an associated marker for AMD in Caucasians but not in Asians. This SNP may have only a small effect on AMD susceptibility. Further studies in larger samples are warranted to confirm its role.Association of a polymorphism rs3775291 in the  Age-related macular degeneration (AMD) is a common degenerative disease among the elderly population, leading to distorted central vision in the early stage and severe visual loss in the advanced stage. The prevalence of early AMD is about 8.01% and advanced AMD 0.37%age-related maculopathy susceptibility 2 (ARMS2) and complement factor H (CFH) are the major associated genes for AMD. However, the ARMS2 risk allele seemed to be more associated with nAMD, whereas the CFH risk allele was preferentially associated with GAAMD is a complex disease with multiple genetic and environmental factors. A recent multi-cohort genome-wide association study (GWAS) has identified 19 AMD loci, accounting for approximately 15–65% of total genetic contribution to AMDTLR3 gene reduced the dsRNA-induced cell death of the RPE cells in vitro and in vivo and had shown potential clinical significanceTLR3 rs3775291 with different subtypes of AMD.It has been reported that the inflammatory cascades are involved in the pathogenesis of AMD6TLR3 rs3775291 was first found in Americans of European descent to protect against GA but not affect CNV or early AMD11131415TLR3 SNPs, including rs3775291 and others, on different subtypes of AMD.17181917811121314151619The main characteristics of the included studies are shown in 813TLR3 rs3775291 and all forms of AMD All these 25 case-control cohorts have been investigated for the association between s of AMD . Of noteGenotype data of rs3775291 was available from 6 studies with 19 cohorts for GA811121315128111315TLR3 rs3775291 was reported in 6 studies with totally 11 cohorts812131415Two studies investigated the association of rs5743303 with AMD in a total of 1695 cases and 1181 controls. Rs5743312 was tested by two studies in 1691 cases and 1181 controls. However, no significant association was detected for these 2 SNPs with AMD in any genetic models in this meta-analysis (data not shown).et al.In the sensitivity analysis, the ORs for any genetic models were not substantively changed after removing the study deviated from HWETLR3 rs3775291 with AMD. We found that this SNP was associated with combined AMD, GA and nAMD in the homozygous and/or recessive models, with the TT genotype showing a protective effect. In the subgroup analysis by ethnicity, rs3775291 showed a suggestive association with combined AMD , GA  and nAMD  in Caucasians, but not in East Asians. However, the associations could not withstand Bonferroni correction. Therefore, current data in the literature provides only suggestive evidence to support the role of TLR3 rs3775291 in AMD susceptibility. Further studies in larger cohorts are needed.In this systematic review and meta-analysis, we have summarized the reported associations of TLR3 gene is located on chromosome 4q35, encoding 904 amino acids. The TLR3 polypeptide contains two characteristic Toll motifs: one is an extracellular leucine rich repeat domain and the other a cytoplasmic interleukin-1 receptor like region222526The TLR3 polymorphisms and their roles in the pathogenesis of AMD have not been well defined as reflected by inconsistent results among different populations. In the study by Yang et al., rs3775291(T) conferred a protective effect against GA (P = 1.24 × 10−7) but not CNV or early AMD in Caucasianset al. reported no evidence of protection of rs3775291 in GA, with an opposite effect size et al., with the exception of three cohorts11121619−5. Therefore, results from multiple cohorts should be combined, e.g., by a meta-analysis, to validate the initial signals identified by candidate gene analyses. As summarized in our meta-analysis, suggestive genotypic (homozygous and recessive) associations were found in rs3775291 for all AMD, GA and nAMD, with the TT genotype showing a protective effect. Of note, leave-one-out sensitivity analysis indicated these significant associations were of no substantive change after removing any individual study (P < 0.05), with one exceptionet al. might have played a predominant role in the current meta-analysis. However, as shown in the forest plots, the odds ratios of the TT genotype were towards the same direction in a majority of study cohorts, suggesting a same trend of effect conferred by the SNP. Nevertheless, the pooled-P values were at borderline, making it difficult to conclude on the association between TLR3 and AMD. Further replication studies in more study cohorts with large sample sizes are needed. In addition, TLR3 rs3775291 conferred protection against all AMD, GA and nAMD in Caucasians but not in Asians, indicating population-specific effect of rs3775291 in Caucasians. Moreover, in a previous study, we meta-analyzed the associations with PCV and found rs3775291 not associated with PCV TLR3 activation may thus be implicated in the development of AMD rather than PCV.The associations of A P = 1.2 × 10−7 bTLR3 rs3775291 in AMD, however, our study revealed several limitations. First, our findings were based on individual unadjusted assessment. Therefore, a more precise analysis should be adjusted for potentially confounding factors if individual data was available. Second, the meta-analysis in Asians, i.e., Chinese and Indians, was performed on a small sample size, involving only 312 cases and 541 controls from 3 studies. There was insufficient power to detect significant association with AMD . Thus, the association of TLR3 rs3775291 with AMD should be elucidated in larger Asian cohorts. Third, although genetic and environmental risk factors are considered as sharing a role to increase the risk of AMD, the included studies provide no gene-gene and gene-environmental interactions. Finally, no replication of the association of rs3775291 with early AMD has been reported since the study of Yang et al.The current meta-analysis is an overview of published genetic reports on TLR3 rs3775291 with AMD in Caucasians but not East Asians, indicating ethnic diversities in the association profiles of TLR3 in AMD. Since TLR3 rs3775291 may confer only a small effect in AMD genetic susceptibility, larger cohorts should be recruited for further replication and validation.In summary, this systematic review and meta-analysis revealed that data in the current literature only provided suggestive evidence to support an association of We used the Preferred Reporting Items for Systematic Review and Meta-Analysis statement (PRIMSA) for this systematic review and meta-analysis.TLR3 in the PubMed, EMBASE and Web of Science databases. All related studies reported up to March 10, 2015 were retrieved without language restriction. We used the following key words: (“toll-like receptor 3” or “toll like receptor 3” or “TLR3”) and (“age-related macular degeneration” or “AMD” or “ARMD” or “age-related macular disease” or “age-related maculopathy” or “ARM”). Furthermore, we also reviewed the reference lists of all retrieved studies to identify eligible studies. To maximize the usable data we also searched in all reported GWAS of AMD including their Literature search was independently performed by two investigators (L.M. and F.Y.T.) for genetic studies on TLR3 and AMD; (3) raw data of allele or genotype counts available; (4) AMD was clearly defined by complete ophthalmic examinations, including fundus fluorescein angiography; (5) using elderly subjects without any other major eye diseases as controls; (6) for reports published by the same study group, the latest study or the one with the largest sample size was included for meta-analysis to avoid double counting. Reviews, case reports, conference reports, treatment response studies, editorial comments and reports without sufficient data were excluded.Eligible studies must meet the following criteria: (1) case-control studies; (2) exploration of association between Two reviewers (L.M. and F.Y.T.) independently extracted the following information from each eligible study: name of first author, publication year, ethnicity, study design, diagnostic criteria of AMD, number of cases and controls, disease subtype, gender composition, mean age, allele and genotype distribution in the case and control groups, and Hardy-Weinberg equilibrium (HWE) test results in controls. Any discrepancies were resolved by thorough discussion with another reviewer (L.J.C.).I2 statistics was used to estimate the degree of heterogeneity among the studies. Summary ORs and 95% CIs were calculated only by the random-effect model to yield more conservative CIs3032We calculated the HWE with genotypes in the control group using Stata , if HWE was not given in the original study. Meta-analysis for each polymorphism was assessed if it had been reported in ≥2 studies. The association was performed using 5 genetic models, including allelic, dominant, recessive, heterozygous and homozygous models. The strength of association was estimated by calculating the summary ORs and 95% CIs. With respect to heterogeneity, the How to cite this article: Ma, L. et al. Association of toll-like receptor 3 polymorphism rs3775291 with age-related macular degeneration: a systematic review and meta-analysis. Sci. Rep. 6, 19718; doi: 10.1038/srep19718 (2016).


CDKN2A gene, is a tumor suppressor that has been widely studied in cancer research. However, the relationship of p16 with prognostic and clinicopathological parameters in patients with bladder cancer remains unclear. Data inclusion criteria were articles reporting on the relationship between p16 expression and the prognosis or clinicopathology in patients with bladder cancer. Meta-analyses were performed with Stata software. Hazard ratios (HRs) or odds ratios (ORs) and 95% confidence intervals (95% CI) were calculated to evaluate the relative risks. The source of heterogeneity was analyzed by subgroup analysis. A total of 37 studies with 2246 cases were included and analyzed. The results identified an important link between downregulated p16 expression and poor prognosis in patients with bladder cancer in terms of recurrence-free survival (RFS), overall survival (OS), progression-free survival (PFS), and some clinicopathological parameters including clinical staging, pathological degree, and lymph node metastasis. Subgroup analysis also showed that low p16 expression could function as a warning sign for RFS and PFS in patients with early-stage (Ta–T1) bladder cancer. In conclusion, p16 might play an essential role in the deterioration of bladder cancer and could serve as a biomarker for the prediction for patients' progression and prognosis. p16, encoded by the These results suggest that abnormal expression of the p16 gene in cells might be associated with tumorigenesis [Bladder cancer is the most frequent malignancy of the urinary tract and the ninth most common cancer worldwide . About 9igenesis .Numerous studies to date have explored the clinicopathological and prognostic significance of p16 in patients with bladder cancer. However, as a result of differences in sample sizes, accuracies of the statistical data, study populations, and interventions, the results remain inconclusive, and evidence-based confirmation by large-scale clinical trials is still lacking. We therefore conducted an in-depth systematic review and meta-analysis to investigate the correlation between abnormal expression of p16 and clinicopathological features, as well as prognosis in patients with bladder cancer. The specific mechanisms are shown in ∗,” “surviv∗,” “follow-up,” “mortality,” “predict,” “course,” “outcome,” and “clinicopathological” were used to search the following domestic and international databases: PubMed, Wiley Online Library, Embase, Cochrane Central Register of Controlled Trials, Science Direct, EBSCO, Google Scholar, Ovid, LILACS, China National Knowledge Infrastructure (CNKI), China Biology Medicine disc (CBMdisc), CQVIP, and Wan Fang, with unified retrieval rules such as Boolean logic. The obtained search results were then analyzed, evaluated, reviewed, and manually screened to determine their relevance.The terms and combinations including “Cyclin Dependent Kinase Inhibitor p16,” “CDKN2A Protein,” “p16INK4A Protein,” “MTS1 Protein,” “Cyclin Dependent Kinase Inhibitor 2A,” “Multiple Tumor Suppressor 1,” “Cdk4 Associated Protein p16,” “TP16,” and “urinary bladder neoplasms,” “bladder tumors,” “bladder cancers,” “bladder carcinomas,” and “prognosInclusion criteria were as follows: (1) patients diagnosed with bladder cancer; (2) immunohistochemical (IHC) detection of p16 expression levels in the tissues; (3) relationships between abnormal expression of p16 and prognostic indicators such as recurrence-free survival (RFS), progression-free survival (PFS), and overall survival (OS) or associations between p16 and clinicopathological features that were evaluated; (4) hazard ratio (HR), odds ratio (OR), relative risk (RR), and 95% confidence intervals (CI) that could be obtained directly from the full article or indirectly calculated with relevant software based on the data provided in the graphics and tables; (5) only the newest studies or the ones with higher quality were retained if the data were repeated in different studies; and (6) studies in English or Chinese.n ≤ 10) and those with no directly or indirectly extractable HR, OR, and 95% CI data; and (3) articles that could not be understood because of language barriers.Exclusion criteria were as follows: (1) cell or animal studies, case reports, letters, reviews, and meta-analyses; (2) articles with similar content or using the same data or those with small sample sizes (Two independent investigators (Xiaoning Gan and Rongquan He) reviewed the articles that met the criteria and extracted data on author, year of publication, nationality, sample size, patient age, detection method of p16, antibody source and dilution, clinical stage, pathological degree, other costudied prognosis-associated genes, cut-off value, outcome, and extraction method of the study subjects. Discrepancies between the two independent investigators in terms of data extraction were resolved by discussion among all the authors.Effects of p16 on the related prognostic indexes were detected by merging the HRs and 95% CI of the included literatures, which were evaluated through the Forest plot and related parameters after the merging. The HRs and 95% CI values mainly came from direct extraction of the original text or survival curve through extraction and calculation by software.The relationships between p16 and the clinicopathological parameters were derived from the binary variable data extracted from the original articles. ORs and 95% CI values came from the binary variable data calculated by Stata software. The data were then combined, and their statistical significance was evaluated by Forest plot and related parameters, to clarify the relationship between p16 low-expression and clinicopathological parameters.Q statistics as follows: no heterogeneity: 0 < I2 < 25%; low heterogeneity: 25% ≤ I2 < 50%; moderate heterogeneity: 50% ≤ I2 < 75%; high heterogeneity: 75% ≤ I2 ≤ 100%. If I2 < 50% and P > 0.10, a fixed-effect model would be used in combination with HRs, ORs, and 95% CI; if I2 ≥ 50% and P ≤ 0.10, then a random-effect model would be selected. Heterogeneity analysis was performed to assess the accuracy of the data, and subgroup and sensitivity analyses were carried out based on professional knowledge.Heterogeneity was measured by P value < 0.05 was considered to indicate statistical significance. Statistical analyses were carried out with StataSE 12.0, Engauge, Photoshop CS5, and Microsoft Office 2007.Publication bias was detected by Begg's funnel plot and Egger's test with Stata software. A two-sided A total of 364 articles were identified from the databases, including 190 English and 174 Chinese articles, 222 of which were excluded because of discrepancies between the study theme and their abstracts. The full text of the remaining 142 articles was then reviewed for their fit with the current study, after which a further 105 articles were excluded because they met one or more of the exclusion criteria, such as the cell or animal studies, reviews, and letters and studies with identical data and no extractable HR, OR, and 95% CI data from the full text or language barrier. The remaining 37 articles –42 with The basic features of the included studies were presented in P < 0.001), with low observed heterogeneity   .Cumulative meta-analysis based on year of publication and sample size demonstrated that the results tended to stabilize with increasing sample size, but there was no obvious relationship between the results and year of publication.Based on sensitivity analysis, the study by Yang et al.  was initP = 0.002) and in Europe . The results of American studies  need to be confirmed with larger sample sizes. The heterogeneity of Asian studies  was lower than the overall heterogeneity, while that of Europe  was higher, calculated with the random-effect model.Subgroup analysis based on geographic region showed that low expression of p16 was associated with RFS in patients with bladder cancer both in Asia .Subgroup analysis based on clinical stage suggested that the effect of p16 on RFS was associated with clinical stage , G1–G3 , and G2-G3 , indicating that the effects of p16 on RFS in patients with bladder cancer were closely associated with pathological grade.Subgroup analysis based on histopathological grade showed that heterogeneity decreased from G1-G2 .Subgroup analysis also showed an effect of cut-off value on the influence of p16 on RFS .In addition, subgroup analysis of early-stage data from 430 subjects from eight studies also demonstrated that low expression of p16 significantly affected RFS in patients with early-stage (Ta–T1) bladder cancer , with no significant observed heterogeneity   .Cumulative meta-analysis and sensitivity analysis indicated relatively low overall heterogeneity and no study with high sensitivity.P = 0.065; I2 = 0%, P = 0.703) and Europe .Subgroup analysis based on geographic area showed a subtle distinction between p16 expression and OS in patients with bladder cancer in Asia .Subgroup analysis was also performed based on clinicopathological stages. However, limitations of sample size led to the impossibility of determining if the effects of p16 expression on OS were associated with these parameters in patients with bladder cancer .Subgroup analysis based on cut-off value indicated that the effects of p16 on OS in patients with bladder cancer were associated with cut-off value , with low heterogeneity detected .A total of 470 subjects in seven studies were included in the ultimate analysis of PFS , 41, 42.Cumulative meta-analysis revealed no obvious characteristics because of the limited range of publication dates and the sample sizes.P = 0.015; I2 = 0%, P = 0.487)  .P = 0.002; I2 = 0%, P = 0.484; T2–T4 group: HR = 1.14, 95% CI = 0.39~3.31, and P = 0.810; I2 = 0%, P = 0.484) and geographical location .Despite a reduced sample size, subgroup analysis of the 347 subjects from five studies , 12, 19 P = 0.002; I2 = 0%, P = 0.932; cut-off value > 10%: HR = 0.95, 95% CI = 0.41~2.18, and P = 0.896; I2 = 0%, P = 0.579).Subgroup analysis based on cut-off value demonstrated some relationship between cut-off value and the influence of p16 expression on PFS .The results from 297 subjects with early-stage (Ta–T1) bladder cancer from four studies , 12, 19 P = 0.149; I2 = 0%, P = 0.825).A total of 187 subjects from three studies were included in the DSS/CSS analysis , 17, 18;P < 0.001; I2 = 1.4%, P = 0.440; T1/Ta: OR = 1.55, 95% CI = 0.87~2.76, and P = 0.134; I2 = 40.5%, P = 0.152; G3/G1-2 [P < 0.001; I2 = 0%, P = 0.519; and H/L [P = 0.580; I2 = 61.8%, P = 0.011; because of the high heterogeneity, a random-effects model was therefore applied. Meanwhile, these results demonstrated significant differences in the effects of low p16 expression in patients with bladder cancer between the two WHO clinical pathological grading methods in 1973 and 2004.The relationship between low expression of p16 and clinicopathological parameters , 21–40 w G3/G1-2 : OR = 3.P = 0.006; I2 = 27.2%, P = 0.240. The small sample size caused the impossibility of demonstrating any significant influence of pathological parameters such as muscle invasion, tumor number (multiple/single), and tumor size on the effect of p16 expression , indicating potential publication bias in terms of RFS.Publication bias was detected by Begg's funnel plot and Egger's test . The poip16, also known as tumor suppressor gene I , is located in 9p21 and is composed of two introns and three exons [ee exons . It is aee exons , 46, 47.Pan et al. performed a meta-analysis of the prognostic significance of abnormal p16 and p21 expression in bladder cancer in 2006 . HoweverThe current study systematically analyzed the relationships between p16 expression and prognostic index and clinicopathological parameters in patients with bladder cancer and showed that low expression of p16 was closely correlated with poor prognosis . HoweverThe current study had some limitations. First, tumors are the result of both environmental and genetic factors, and p16 may thus be only one of several factors involved in the whole process of bladder carcinogenesis. Secondly, heterogeneity may result from differences in intervention measures , immunohistochemical techniques , and the HR extraction methods used in the included studies. Finally, the exclusion of articles because of language barriers and of studies that were not published because of a lack of sufficient data may have led to potential publication bias.In conclusion, the results of the current study provide evidence for a relationship between p16 expression and prognosis and clinicopathological features in patients with bladder cancer. The results of this meta-analysis will help to inform about the development of clinical guidelines promoting best medical care for patients with bladder cancer. Further studies are required to investigate the combined influence of genetic and environmental factors on the development and progression of bladder cancer.


CI 1.03‐1.15; I2=33.4%). The positive association persisted in almost all strata of subgroup analyses based on study design, location, study quality, type of controls, and confounder adjustment, although in some strata statistical significance was not detected. By evaluating the number of parity, we identified that both parity number of 2 versus nulliparous and parity number of 3 versus nulliparous demonstrated significant positive associations . The dose‐response analysis suggested neither a non‐linear nor linear relationship between the number of parity and thyroid cancer risk. In conclusion, this meta‐analysis suggests a potential association between parity and risk of thyroid cancer in females. However, the lack of detection of a dose‐response relationship suggests that further studies are needed to better understand the relationship.Although observational studies have assessed the relationship between parity and thyroid cancer risk, the findings are inconsistent. To quantitatively assess the association, we conducted a systematic review and meta‐analysis. PubMed and Embase were searched up to January 2015. Prospective or case–control studies that evaluated the association between parity and thyroid cancer risk were included. We used the fixed‐effects model to pool risk estimates. After literature search, 10 prospective studies, 12 case‐control studies and 1 pooled analysis of 14 case‐control studies including 8860 patients were identified. The studies had fair methodological quality. Pooled analysis suggested that there was a significant association between parity and risk of thyroid cancer (RR for parous versus nulliparous: 1.09, 95%  As the most common type of the endocrine malignancies, thyroid cancer causes a large number of deaths that is higher than the combined number of all other endocrine cancers This meta‐analysis was performed in accordance with the MOOSE guideline A search of PubMed (MEDLINE) and Embase was conducted from each database's inception to January 2015 for studies of humans published in English. We used the following search keywords and Medical Subject Heading terms: (parity OR pregnancy OR livebirth OR reproductive OR reproduction OR reproductive factors) AND (papillary OR follicular OR thyroid) AND (cancer OR neoplasm OR carcinoma OR tumor OR adenoma OR cancers OR neoplasms OR carcinomas OR tumors OR adenomas). We also reviewed references of relevant review articles to identify additional potential studies.Studies were eligible if they (1) were case–control studies or prospective studies; (2) evaluated the association between parity and risk of thyroid cancer; (3) presented odds ratio (OR), relative risk (RR), or hazard ratio (HR) estimates with 95% confidence intervals (CI) or data necessary to calculate them. Studies were excluded if they used a cross‐sectional study design. Studies primarily focusing on subjects with extensive exposure to radiation were not included because exposure to radiation is the most well‐established risk factor for thyroid cancer, and it would make the studied population significantly different from more general population and might induce bias for the research question of interest. Studies were included regardless of publication status, sample size and length of follow‐up. If multiple publications from the same study were identified, we included the study with the largest number of cases and most relevant information, like previous studies A pair of investigators independently carried out the abstract screening, full‐text screening, data extraction, and quality assessment. Disagreements were resolved by consensus. Data extracted from each study included: the first author's last name, year of publication, study region, study design, characteristics of study population . If multiple estimates of the association for the same outcome were reported, we extracted the estimate that adjusted for the most appropriate covariates, like previous studies To assess the study quality, we used the Newcastle–Ottawa Quality Assessment Scale I2 to assess the heterogeneity across the included studies, where I2 > 50% suggests substantial heterogeneity The RRs and corresponding 95% CIs from each of the included studies were used as the measure of association across studies. Due to the rarity of thyroid cancer, ORs and HRs were deemed equivalent to RRs and we used RRs to represent measures. We used the For the dose–response analysis, we explored potential nonlinear and linear relationship between the number of parity and risk of thyroid cancer P‐value of 0.05 was used as the threshold for determining significant publication bias. All statistical analyses were performed with Stata .Publication bias was evaluated via Egger's test The detailed steps of the literature search were shown in Figure I2 = 33.4%; Table P for bias: 0.878) and Begg's test (P for bias: 1.000). Sensitivity analysis revealed that the 23 study‐specific RRs of parous versus nulliparous ranged from a low of 1.07  after omission of the study by Negri et al. I2 = 32.4%) after omission of the study by Rossing et al. After pooling results from all available studies, there was a significant positive association between risk of thyroid cancer and parity for parous versus nulliparous , with no considerable heterogeneity (I2 = 3.6%), although the association was not statistically significant. On the other hand, both parity number of two versus nulliparous and parity number of three versus nulliparous demonstrated significant positive association with the risk of thyroid cancer .We assessed the associations between different number of parity  and risk of thyroid cancer, respectively Table . Parity P = 0.69 for the linear trend), with significant heterogeneity (P for heterogeneity: <0.0001). There seemed not be a clear dose–response relationship between the number of parity and thyroid cancer risk.Based on the dose–response analysis, we did not detect a nonlinear dose–response relationship between the number of parity and risk of thyroid cancer. Assuming a linear dose–response relationship, the combined RR per live birth was 1.01  demonstrated that such a significant positive association with thyroid cancer risk persisted for both parity number of two versus nulliparous and parity number of three versus nulliparous. However, the dose–response analysis did not suggest a significant nonlinear or linear relationship between the number of parity and thyroid cancer risk. Overall, these findings suggested that parity might be associated with risk of thyroid cancer in females, while the exact relationship needs exploration and clarification in further studies.Although the exact biological mechanism underlying the potential association between parity and risk of thyroid cancer has not been completely established, plausible explanations have been suggested by basic research. During pregnancy estrogens are elevated, which potentially influence the proliferation as well as enhance the adhesion, migration, and invasiveness of malignant thyroid cells In the subgroup, analyses of the association between thyroid cancer risk and parity for parous versus nulliparous, significant positive association was also detected in subgroups of studies with a case–control design, case–control studies with hospital‐based controls, high–quality studies, and studies with confounder adjustments. We acknowledge that studies with a case–control design are more susceptible to bias compared with studies with a prospective design. Similarly, case–control studies with hospital‐based controls may be more susceptible to bias compared with those with population‐based controls. On the other hand, even though the detected associations in many other subgroups did not reach statistical significance, the directions of the associations tend to be positive. The trend of a positive association was also suggested for parity number of one versus nulliparous. These suggest that the detected positive association between parity and thyroid cancer risk may be real and warrants further clarification.Several reasons may explain the inconsistencies of the association between parity and thyroid cancer across included studies. For example, not all included studies sufficiently adjust for relevant covariates. Besides parity, several other reproductive factors like age at first pregnancy, OC use, and age at menopause are suggested to influence thyroid cancer risk as well Our study has several strengths. To the best of our knowledge, this is the most comprehensive meta‐analysis evaluating the association between parity and thyroid cancer risk. A systematic review previously assessed the association Several potential limitations must be acknowledged for the interpretation of our findings. First, we did not have access to the individualized primary data from the included studies, and the risk estimates used in pooling might not be fully adjusted for. Relevant covariates including age, BMI, iodine intake, use of OC, HRT, and fertility treatment were not always adjusted for in the included studies. Residual confounding may thus be an issue for our findings. Further well‐designed studies with full adjustments are needed. Second, during the dose–response analysis, the highest levels of number of parity in different studies have wide range of values, which may cause the exposure values to not be accurately assigned. This may be one reason that we did not detect a linear or nonlinear dose–response relationship between the number of parity and risk of thyroid cancer, which seemed to be suggested based on the increasing risks over parity of 1, 2, and 3 (Table Based on a summarization of relevant evidence from epidemiological studies, parous versus nulliparous was positively associated with risk of thyroid cancer. A similar positive association was also detected for both parity number of two versus nulliparous and parity number of three versus nulliparous. However, no linear or nonlinear relationship between the number of parity and thyroid cancer risk was detected. Although parity might be associated with the risk of thyroid cancer in females, further studies are warranted to better clarify the relationship.There are no competing interests to declare.


PARP-1 rs1136410: C>T is among the most studied polymorphisms and likely involved in human carcinogenesis. However, results from previous studies are inconclusive. Thus, a meta-analysis was conducted to derive a more precise estimation of the effects of this enzyme.Poly(ADP-ribose) polymerase-1 (PARP-1) is a nuclear chromatin-associated enzyme involved in several important cellular processes, particularly in the DNA repair system. PARP-1 Val762Ala polymorphism and cancer risk was found when all of the studies were pooled into the analysis . The subgroup analysis of cancer types revealed that the –762Ala allele was associated with increased risk of gastric, cervical, and lung cancers and a decreased risk of glioma. In addition, a significantly increased risk of cancer associated with the polymorphism was observed in Asian descendents . These results also indicated that a joint effect between PARP-1 Val762Ala and XRCC1 Arg399Gln could be involved in the risk of cancer development .A comprehensive search was conducted in the PubMed and EMBASE databases until December 9, 2013. A total of 39 studies with 16,783 cancer cases and 23,063 control subjects were included in the meta-analysis on the basis of the inclusion and exclusion criteria. No significant association between the PARP-1 Val762Ala may be involved in cancer development at least in some ethnic groups (Asian) or some specific cancer types .The present meta-analysis provides evidence that the  The etiology and development of cancer are a result of complex interactions between genetic and environmental factors. Physical and chemical agents originated from either endogenous processes, such as cellular metabolism, or exogenous exposure, including ionizing radiation, tobacco smoke, and genotoxic chemicals, are responsible for oxidative cell DNA damage; when left unrepaired or incorrectly repaired, cell DNA damage may lead to mutations and genomic instability Poly(ADP-ribose) polymerase-1 (PARP-1), also called adenosine diphosphate ribosyl transferase, is one of the most important components of the BER system. PARP1 is a nuclear nick sensor enzyme that becomes activated in response to DNA breakage PARP1 gene, located on chromosome 1q41–42, is approximately 47.3 kb in length and consists of 23 exons. Numerous single nucleotide polymorphisms (SNPs), including 17 non-synonymous SNPs, have been identified in PARP-1; among these SNPs, rs1136410 at codon 762 in exon 17, a non-synonymous T→C polymorphism changing valine to alanine, is the most extensively investigated. This polymorphism is located in the sixth helix of the COOH-terminal NAD-binding region with all of the catalytic activities of the full-length enzyme. This amino acid change contributes to low poly(ADP-ribosyl)ation activities in a dosage-dependent manner, thereby impairing DNA repair and enhancing the susceptibility of variant allele carriers to damage caused by environmental carcinogens and cancer risk PARP-1 Val762Ala polymorphism and susceptibility to cancers have been discussed PARP-1 in cancer. This meta-analysis also aimed to quantify the potential of heterogeneity between studies.The human Relevant publications were identified by conducting a literature search in PubMed and EMBASE databases using the following search terms: PARP-1 or ADPR, variant or polymorphism or SNP, and cancer or carcinoma or tumor. The last search was updated on December 9, 2013. The references of the identified studies and reviews were also screened to find additional eligible studies. If studies with overlapped subjects were reported, only the one with the most complete data was included in the meta-analysis. Search results were limited to studies published in English.PARP-1 Val762Ala polymorphism and cancer susceptibility; and (3) sufficient genotype data were provided to estimate the odds ratio (OR) and a corresponding 95% confidence interval (CI). Studies were excluded if the following criteria were satisfied: (1) case-only, case reports, or reviews; (2) duplicate of previous publications; (3) family-based studies; and (4) based on insufficient data for calculation.Studies were included in our meta-analysis if the following criteria were satisfied: (1) studies were designed as cohort or case-control; (2) studies investigated the association between Two investigators dependently reviewed the publications and obtained information according to a standard data form. The following data were extracted from each study: name of first author; year of publication; country or region of origin; ethnicity of the study population; cancer type; number of cases and controls; allele and genotype frequency; evidence of Hardy-Weinberg equilibrium (HWE) in controls; source of controls; and genotyping method. Disagreements between the two investigators were resolved by discussing the results with a third investigator.PARP-1 Val762Ala polymorphism and the risk of cancer was measured by OR with 95% CI in five genetic models, including dominant model (VA + AA vs. VV), recessive model (AA vs. VA + VV), homozygous model (AA vs. VV), heterozygous model (VA vs. VV), and allele model (A vs. V). The significance of the pooled OR was determined by a Z-test, and P<0.05 was considered statistically significant. A statistical test to determine heterogeneity between studies was performed using Q-test and 2I test. In the Q-test, P>0.10 indicates the absence of heterogeneity. The pooled OR estimates of each study were calculated using the fixed-effect model, the Mantel-Haenszel method. Otherwise, a random-effect model, the Dersimonain and Laird method, was applied. The 2I test was used to quantify the effect of heterogeneity (ranges from 0% to 100%); The test represents the proportion of inter-study variability that can be attributed to heterogeneity rather than by chance. Subgroup analyses were also performed to evaluate the potential effects of ethnicity, cancer types, source of controls, and genotyping method. Sensitivity analysis was conducted by omitting each study to identify the effect of an individual study on the pooled OR. Publication bias was qualitatively detected using Begger's funnel plots, and Egger's linear regression test was performed to determine funnel plot asymmetry . All of the P values were two-tailed. Statistical analyses were performed using STATA version 11.0 .The strength of the association between the A total of 84 articles relevant to search keywords were identified after our literature search from PubMed and EMBASE was completed. According to the inclusion criteria, 45 studies were excluded. Among these studies, two were excluded because of a lack of genotyping data PARP-1 V762A and cancer risk are summarized in The meta-analysis findings of the correlation between PARP-1 V762A polymorphism . To explore the source of heterogeneity, we performed stratified analyses on ethnicity, cancer type, source of controls, and genotyping method. In the subgroup analysis of ethnicity, PARP-1 V762A was significantly associated with an increased risk of cancer in Asian populations in all of the genetic models  and gastric cancer . By contrast, the PARP-1 V762A polymorphism was significantly associated with a decreased risk of glioma in three genetic models .Considering that PARP-1 functionally interacts with XRCC1 in BER processes, we performed a gene-gene interaction analysis of the five studies that reported joint effects between Sensitivity analysis was conducted to verify the effect of each study on the overall OR by repeating the meta-analysis, but any single study was omitted at each time. In t = −0.11, P = 0.916).Begger's funnel plot and Egger's test were performed to evaluate the publication bias of the studies. The shape of the funnel plots showed that the dots were nearly symmetrically distributed predominantly in pseudo 95% confidence limits ation and trans-poly(ADP-ribosyl)ation of histone H1. Lockett et al. PARP-1 Val762Ala polymorphism reduces the enzymatic activity of PARP1 in response to oxidative damage. Molecular epidemic studies have also been conducted to investigate the functional relevance of this variant with susceptibility to cancer. However, results remain inconsistent.Thus far, approximately 1,066 single-nucleotide polymorphisms in the PARP-1 Val762Ala polymorphism with overall cancer risk. In the stratified analysis by ethnicity, the variant –762Ala allele was significantly associated with an increased cancer risk among Asian populations. By contrast, no significant correlation was detected among Caucasians. The discrepancy in ethnicity could be attributed to the evident difference in the minor allele frequency (MAF) of Val762Ala polymorphism in Asians and Caucasians in our meta-analysis . This genetic polymorphism variance with ethnicity was consistent with those described in a previous study PARP-1 variant genotypes may possibly be tissue specific because of high or low PARP-1 expression levels in different tumor tissues ERCC1 –399Gln and PARP-1–762Ala on increased cancer risk in a homozygous genetic model. However, this result should be carefully interpreted because of a relatively small sample size; moreover, this result should be confirmed by conducting further analysis of additional published studies.A total of 39 studies with 16,783 cancer cases and 23,063 controls were considered in the present meta-analysis. The results indicated no significant association of PARP-1 Val762Ala in increased cancer risk among Asian populations. Furthermore, cancer types in the study were more multifarious (seven types) and a significant association was found in cervical, lung, and gastric cancers, as well as glioma. In addition, the potential interaction effect of XRCC1 Arg399Gln on PARP-1 Val762Ala was also evaluated in the present analysis.Compared with two previous meta-analyses, our meta-analysis involved a remarkably larger number of studies (39 vs. 21 and 28) and provided a more comprehensive and reliable conclusion. Pooling the data from 39 studies, we reconfirmed the function of PARP-1 Val762Ala and XRCC1 Arg399Gln genotypes on the risk of cancer was addressed in the present study. However, the lack of individual data from the included studies limited the further evaluation of other potential interactions, as in other genes and environment factors. For instance, only two studies have reported the combined effect of XRCC1 Arg194Trp and PARP-1 Val762Ala genotypes on the risk of cancer Some potential limitations of this study should also be considered. First, the pooled results were based on unadjusted estimates because not all of the studies provided adjusted ORs; when these studies revealed adjusted ORs, these ORs were not adjusted by the same confounders. Hence, a precise analysis should be performed if individual data, such as age, sex, BMI, and smoking and drinking status, were available. Second, several factors, such as gene-gene or gene-environment interaction, may influence gene-disease factor. The joint effect between PARP-1 Val762Ala with increased cancer risk among Asian populations. The same results were observed in the subgroups of gastric, cervical, and lung cancers, as well as in studies using PCR-RFLP genotyping method. Our findings suggested that the PARP-1 Val762Ala polymorphism may function in cancer development in an ethnicity- or cancer-specific manner. Well-designed epidemiological studies should be conducted by carefully matching cases and control subjects to verify our observations. Further studies may focus on the influence of gene-gene and gene-environment interactions on the association of cancer and PARP-1 Val762Ala polymorphism.In conclusion, the present meta-analysis provided strong evidence of the association of Checklist S1PRISMA checklist.(DOC)Click here for additional data file.


Background: Esophageal cancer (EC) and gastric cancer (GC) are common cancers and leading causes of cancer deaths worldwide. Many studies have investigated the association between dietary flavonols intake and the risk of EC and GC, but the results are inconsistent. Hence, we conducted a systematic analysis of relevant population-based studies to assess the association and derive a more precise estimation. Methods: The Cochrane, PubMed and Embase databases were searched to identify articles published through January 2016 that met the predetermined inclusion criterion. Twelve studies involving 4593 patients and 519,378 controls were included. Results: The summary odds ratios (ORs) of EC, GC and the two combined were respectively 0.88 (95% CI: 0.73–1.08), 0.80 (95% CI: 0.70–0.91) and 0.83 (95% CI: 0.74–0.92) for the highest category of dietary flavonols intake compared with the lowest. No significant heterogeneities were observed in these studies. Further analysis showed that the pooled ORs of EC and GC for cohort, population-based case-control and hospital-based case-control studies were 0.90 (95% CI: 0.61–1.34), 0.92 (95% CI: 0.72–1.18), 0.68 (95% CI: 0.38–1.24) and 0.83 (95% CI: 0.65–1.06), 0.84 (95% CI: 0.45–1.59), 0.70 (95% CI: 0.56–0.88). The subgroup analyses revealed a significant association of flavonol intake with a reduced risk of noncardia gastric adenocarcinoma but not gastric cardia adenocarcinoma. Moreover, significant inverse associations of flavonol intake with GC risk were observed in women but not in men, in smokers but not in nonsmokers, in European populations but not in American populations. Similarly, a significant inverse association of flavonols intake with EC risk was also observed in smokers but not in nonsmokers. Conclusion: High intake of dietary flavonols is significantly related to a reduced risk of GC, especially in women and smokers. Esophageal cancer (EC) and gastric cancer (GC) are, respectively, the eighth and the fourth most common cancer and leading causes of cancer deaths worldwide ,2. It isin vitro anti-proliferative and pro-apoptotic agents in a series of tumor cells. Quercetin and myricetin can impact the etiology of certain diseases ranging from oxidative stress and inflammation to carcinogenesis and cancer progression [High fruit and vegetable consumptions are associated with beneficial health effects, and these effects have been partly attributed to flavonoids, a group of polyphenolic compounds, which occur ubiquitously in plant foods. Dietary flavonols, mainly including quercetin, myricetin and kaempferol, are a subclass of flavonoids in a daily diet. Dietary flavonols mostly exist in black tea, onions, broccoli, beer, apples, wine and mixed salads in western country diets ,10. In rgression ,12. EpidA systematic search of the literature published through 2 January 2016, was conducted using the Cochrane, PubMed and Embase databases. The following search terms were used: “flavonoid”, “flavonoids”, “flavonols”, “flavonol”, “quercetin”, “myricetin”, “kaempferol”, “gastric cancer”, “stomach cancer”, “esophagus cancer” and “oesophageal cancer” ((Flavonoid OR flavonoids OR flavonols OR flavonol OR quercetin OR kaempferol OR myricetin) AND ). We also performed a manual search via reference lists. Only full-length journal articles with a prospective cohort, population-base case-control (PBCC) or hospital-based case-control (HBCC) study design were considered.Articles were eligible for the present meta-analysis if they conformed to the following criteria: (i) the study design was a population-based study, including a cohort, PBCC or HBCC study; (ii) a relatively complete assessment of dietary flavonols intake was performed; (iii) the association of dietary flavonols intake with the risk of EC or GC was specifically evaluated; and (iv) the relative risk (RR), hazard ratio (HR), or odds ratio (OR) and the corresponding 95% confidence interval (95% CI) values were available. The data from each study and article fulfilling the inclusion criteria were extracted carefully by two independent reviewers. The following information from each article was recorded: (i) the first author’s name and publication year; (ii) the country of origin, the follow-up time used in the study and the study design ; (iii) the population (numbers of cases and controls); (iv) baseline intake of flavonols ; (v) dietary flavonols included in the dietary assessment; (vi) the level of dietary flavonols intake (mg/day); (vii) the RR or OR values from the most fully adjusted model and their 95% CI values; (viii) the listed confounders adjusted for the multivariate analysis. In addition, due to the low incidence of EC and GC, the OR was assumed to be the same as the RR, and the summary results were reported as OR for simplicity .I2 statistics. Heterogeneity is considered to be significant for p < 0.10 or I2 > 50%. Summary ORs calculations used the Mantel-Haenszel fixed-effects model [p < 0.10). All of the analyses were performed using STATA version 11.0 . A p value < 0.05 was considered to be statistically significant unless otherwise specified. The summary ORs and corresponding 95% CIs of the included studies were used as a measure to assess the association of dietary flavonols intake and the risk of EC and GC. We estimated between-study heterogeneity in each meta-analysis by the Q test and ts model  and the ts model ,16. In tThe present systematic search of the related literatures identified a total of 579 studies. The characteristics of the 12 included studies are shown in Among these 12 studies, two were conducted in the USA, one was conducted in Korea, and the other nine were conducted in Europe . One study only adjusted for age and dietary supplements, and the other 11 studies all adjusted for a wide range of potential confounders of cancer, including age, sex, total energy intake, physical activity, body mass index, smoking status, educational level, alcohol intake, fruit and vegetables consumption, and red and processed meat consumption.We calculated the summary OR values of EC and GC from the combination of the included studies. Forest plots in Furthermore, we conducted subgroup analyses by study design, cancer subtype, sex, smoking, population and publication time. As shown in No statistical heterogeneities were found across the studies in the overall pooled estimate. Sensitivity analysis was then performed to evaluate the stability of the results, in which each individual study was sequentially dropped. The analysis was conducted by excluding any single study in turn and pooling the OR of the remaining studies. The summary ORs did not substantially change, with a range from 0.85 (95% CI: 0.68–1.06) to 0.91 (95% CI: 0.74–1.13) for EC and 0.73 (95% CI: 0.63–0.84) to 0.84 (95% CI: 0.73–0.97) for GC, for the highest intake category compared with the lowest. Begg’s funnel plots and Egger’s tests were performed to assess the potential publication bias in the included studies. As shown in The WHO has indicated that over the next 20 years there will be an approximately 50% increase in new cancer cases attributed to the steadily increasing proportion of elderly people in the world. If the current smoking levels and the adoption of unhealthy lifestyles persist, the increase will be even greater ,2. ThereThe present meta-analysis supports a significant association of dietary flavonols intake with a reduced risk of GC, as well as EC and GC combined, especially in smokers. However, the interpretation of the results should be cautious. The present study has several potential limitations. The main limitation of this meta-analysis is the small number of perspective cohort studies and EC cases included. Especially the number of included studies for subgroup analysis was largely limited, some summary estimates even could not be calculated in several subgroup analyses due to the limited number of studies. So the results, especially subgroup analysis, should be interpreted with caution. More population-based studies are needed to validate the suggestions of the present meta-analysis. Furthermore, another potential confounding factor is the bioavailability of dietary flavonol compounds when estimating the amount of flavonol intake necessary to reduce the risk of EC and GC. In fact, little is known about flavonol absorption in the gastrointestinal tract; the metabolism of flavonols varies by individual, and the degree to which flavonols might have direct effects on epithelial surfaces as they traverse the esophagus and stomach is unclear.Studies have indicated that dietary flavonol compounds have a variety of bioactivities that may repress carcinogenesis and cancer progression, such as anti-oxidant, anti-inflammation, anti-proliferative and anti-angiogenic activities ,32,33,34We are unaware of any previous sysmatic analysis of the association for dietary flavonols with risk of EC. The present meta-analysis shows that the pooled ORs of EC for the highest and the second highest flavonols intake compared with the lowest were 0.88 (95% CI: 0.73–1.08) and 0.74 (95% CI: 0.61–0.90), respectively. The results suggested that the higher, but not the highest, flavonols intake may be associated with a significant reduced risk of EC. The phenomenon may be attributed to the reason mentioned before. The diet with highest intake of flavonols tends to have high consumption of hot beverages and alcohol, which are risk factors for EC. Therefore, carefully identifying the sources of dietary flavonols and minimizing the effects of confounding factors are important and essential for evaluating the association between flavonols and cancer risks. Woo and Kim conducted a meta-analysis of GC risk and dietary flavonoids previously; a significant association was found only between flavonols and GC risk based on a limited number of selected studies (OR (95%CI) = 0.68 (0.46–0.99)) . The preIn conclusion, the present study indicated a significant association of dietary flavonols intake with a decreased risk of GC, as well as EC and GC combined, especially in smokers. Nevertheless, the results should be interpreted with caution. Residual confounding by alcohol, hot beverages consumption, smoking and total energy intake, etc. is a great concern in studies of EC and GC. Therefore, more carefully designed studies, especially cohort studies, should be conducted to investigate the association of dietary flavonols intake with the risk of EC and GC.


P = 6.77×10-4). Functional analyses on HOTAIR Expression showed that the expression level of HOTAIR in OS tissues was significantly higher than that in corresponding normal tissues, and subjects with the rs7958904 CC genotype had significantly lower HOTAIR RNA levels than those of other genotypes. This should be the first study to examine the association between HOTAIR variants and OS risk.Osteosarcoma (OS) is the most common primary malignant bone tumor in adolescents and young adults. However, the essential mechanisms underlying osteosarcomagenesis remain obscure. The HOTAIR, a well-known long noncoding RNA (lncRNA), is involved in pathogenesis and progress of multiple tumors. To reveal the potential role of lncRNA HOTAIR in OS carcinogenesis, we conducted a two-stage, case-control study among Chinese population with 900 OS cases and 900 controls to evaluated associations of its genetic variants with OS risk. We found that C allele of rs7958904 was associated with a significantly decreased OS risk when compared with G allele (OR: 0.77; 95% CI: 0.67-0.90;  Osteosarcoma (OS), which derives from primitive bone-forming mesenchymal cells, is the most common primary bone malignancy and the eighth-most common form of childhood cancer, comprising 2.4% of all malignancies in pediatric patients, and about 20% of all primary bone cancers -5. AccorRecently, long non-coding RNAs (lncRNAs) have been identified for their wide range of biological regulatory functions in the carcinogenesis and progression of many cancers -14. LossP > 0.05), which indicates that the frequency matching was adequate. The tumors were mainly located at extremities, and half of the cases were metastatic OS. About 39.8% of cases' enneking stage were defined as III in stage 1, and 40.1% in stage 2.In current study, a total of 900 OS cases and 900 controls were recruited in two stages, and there were no significant differences between OS cases and healthy controls for each stage regarding to age and gender . The genotype distribution of all the tagSNPs and their associations with OS risk in our Stage 1 are shown in Table P value = 0.023, 0.024 respectively). For rs7958904, C allele was associated a significantly decreased OS risk when compared with G allele ; while for rs874945, A allele was associated a significantly increased OS risk when compared with G allele . No notable associations between rs4759314 and OS risk in the Stage 1 were observed. Therefore, we validated the effects of rs874945 and rs7958904 in the Stage 2, to further confirm the association observed in the first stage. As shown in Table Ptrend = 0.614). However, the protective effects of rs7958904 still existed in the Stage 2 (Ptrend = 0.011). When combined together, C allele was significantly associated with a decreased OS risk when compared with G allele . The adjusted OR for the carriers with the CG genotype was 0.82 (95% CI: 0.67-1.00) and for those with the CC genotype was 0.57 (95% CI: 0.40-0.81) compared with the GG genotype. We also conducted stratified analyses by age, gender, and tumor location  among the controls were in accordance with Hardy-Weinberg equilibrium . Subjects with the rs7958904 CC genotype had significantly lower HOTAIR RNA levels (mean ± SD) than those with the GG genotypes in normal and OS tissues (P < 0.01).To confirm the abnormal expression of HOTAIR in OS patients, we evaluated the HOTAIR levels in 100 paired tissues of OS patients and corresponding normal tissues. As shown in Figure Deeper understanding of lncRNAs and their role in carcinogenesis could possess a large number of potential clues for developing novel therapeutic agents for OS. In this two-stage, case-control study, we examined the relationship between HOTAIR variants and OS risk among Chinese population. We identified that people with C allele of SNP rs7958904 had a decreased risk of developing OS. SNP rs7958904 also has a genotype-specific effect on lncRNA HOTAIR expression. Our findings support the hypothesis that functional genetic variants influencing lncRNA expression may explain a part of OS genetic basis. To our knowledge, this is the first study to examine the association between HOTAIR variants and OS risk.in vitro and in vivo [lncRNA HOTAIR has been widely explored as a functional lncRNA participating in multiple cancers . In 2007 in vivo . Except  in vivo . Conclusin silico analysis, the secondary structure of HOTAIR was remarkably changed with rs7958904 G/C variant [In current study, we determined C allele of HOTAIR rs7958904 could decrease the OS risk, as well as the expression level of HOTAIR . These results was consistent with a previous study by Xue et al , which i variant . However variant , was notin vivo studies found HOTAIR was notably up-regulated in OS tissues than in adjacent normal tissues, and subjects with the rs7958904 CC genotype had significantly lower HOTAIR RNA levels than other genotypes. Strength of this study includes large sample size, study design, and homogeneous participants. Larger prospective studies and further studies into the detailed biological mechanisms of HOTAIR SNPs are warranted to confirm our results.In conclusion, we identified a SNP located in HOTAIR gene (rs7958904) that was significantly associated with decreased risk of OS in our two-stage, case-control study. Furthermore, Totally included in this study were 500 subjects diagnosed with OS, and 500 healthy controls in stage 1, as well as 400 OS patients and 400 healthy controls in an independent validation stage (stage 2). Pathological diagnosis was made by an experienced pathologist and double-checked by another pathologist. The healthy controls free from any cancer and matched by gender and age, were recruited when they were attending a routine examination. Blood samples (5 ml) were obtained from the subjects who participated in the study. A structured questionnaire was used to elicit detailed information on demographic factors. All specimens were handled and made anonymous according to the ethical and legal standards. This study was approved by the institutional Review Boards. Written informed consents were obtained according to the Declaration of Helsinki from both groups.Genomic DNA was extracted from peripheral blood mononuclear cells using the QIAamp DNA whole blood kit . The tagSNPs were selected basing on data of Han Chinese population (HCB data) of the HapMap Project  covering 10232bp region (6232bp HOTAIR locus and 2kb upstream as well as 2kb downstream regions of the HOTAIR gene). Three htSNPs  were selected with Haploview version 4.2 software. SNP genotyping was performed by the Sequenom MassARRAY RS1000 while Sequenom Typer 4.0 Software was used to perform data management and analysis. The genotyping analysis was done blind as regards participants. To control the quality, the selected PCR-amplified DNA samples were examined by DNA sequencing to confirm genotyping results.−ΔCt method. All assays were conducted by using the ABI 7900 HT system . All reactions were performed in triplicate.A total of 100 OS tissues were obtained from pre-treatment patients. RNA was extracted from the frozen tumor and corresponding normal tissues by standard methods using TRIzol reagent  according to the manufacturer's instructions. The expression of HOTAIR was determined by SYBR Green Assay and the levels was calculated relative to expression of β-actin by the 22 test  or Student's t test (continuous data). Hard-Weinberg analysis was performed by comparing the observed and expected genotype frequencies of HOTAIR using chi-square test. The odds ratios (OR), with 95 % confidence interval (CI), were calculated by an unconditional logistic regression model. We evaluated the log-additive  inheritance models for each SNP in relationship to OS case status. Statistical analysis was performed using SPSS13.0 software package .


TP53 codon 72 polymorphism and oral squamous cell carcinoma (OSCC) susceptibility; however, current results are inconsistent. We therefore performed this meta-analysis to thoroughly investigate any association among Asian patients.Several epidemiological studies have previously investigated the association between the A comprehensive search of PubMed and Embase databases was performed up to December 2013. We only considered studies consisting of patients diagnosed with OSCC by pathological methods. Statistical analyses were performed using Review Manager (RevMan) 5.2 software and odds ratios (ORs) with 95% confidence intervals (CIs) were used to assess the association.TP53 codon 72 polymorphism and OSCC susceptibility . However, subgroup analysis demonstrated an association between the TP53 codon 72 polymorphism and human papillomavirus (HPV)-related OSCC patients. Although statistical heterogeneity was detected, there was no evidence of publication bias.A total of 11 case–control studies involving 2,298 OSCC patients and 2,111 controls were included. We found no association between the TP53 codon 72 polymorphism is not associated with OSCC in Asians without the presence of HPV infection. Further research is necessary to determine if such a relationship exists in HPV-related OSCC patients.Current results suggest that the  Oral cancer is ranked as the 11th most common type of cancer worldwide . Oral sqTP53) gene, located on chromosome 17p13, is one of the most frequently mutated genes in human cancers and has been reported to be a significant determining factor in carcinogenesis . Reference lists of the included studies and published meta-analyses on related topics were also screened for additional studies.Two authors independently extracted the following trial data from included studies: last name of the first author, publication year, countries of origin, HPV status of cases, source of control, number and genotyping distribution of cases and controls, diagnostic method for OSCC, genotyping method, and Hardy-Weinberg Equilibrium (HWE) for controls . DisagreI2 statistic [I2 was more than 40%, we switched to a random-effects model. The odds ratios (ORs) and relevant 95% confidence intervals (CIs) were used to quantify the strength of association between the TP53 codon 72 polymorphism and OSCC susceptibility using five genetic models: Arg vs. Pro, ArgArg vs. ProPro, ArgPro vs. ProPro, (ArgPro + ArgArg) vs. ProPro, and ArgArg vs. (ProPro + ArgPro). Additionally, subgroups analyses based on HPV status, source of controls, and HWE status for controls were performed. Publication bias was detected by examination of funnel plots. All statistical analyses were conducted using Review Manager (RevMan) software (version 5.2 for Windows).We employed the fixed-effect analytical model first to pool results of the included studies, and the tatistic  was usedOur systematic literature search identified 278 studies that met the inclusion criteria. After deduplication and exclusion of the clearly irrelevant studies, we eventually included 11 case–control studies -29 involTP53 codon 72 polymorphism and OSCC susceptibility in Asians [ for Arg vs. Pro;  for ArgArg vs. ProPro;  for ArgPro vs. ProPro, Figure Table TP53 codon 72 polymorphism and HPV infection was observed (Table Results of the subgroup analyses stratified by source of controls and HWE status for controls were similar to those of the overall analyses. However, when stratified by HPV status, a correlation between the A funnel plot based on the ArgPro vs. ProPro genetic model showed a relatively symmetrical distribution, enabling us to conclude that there was no publication bias Figure .TP53 codon 72, and Pro to Arg is the most informative polymorphism in the TP53 gene and have been found to be associated with human cancers [TP53 codon 72 polymorphism and oral cancer susceptibility both reached the same conclusions [TP53 codon 72 polymorphism plays a role in the development of OSCC.Arg and Pro are two distinct functional alleles that are encoded by the  cancers ,30. Amon cancers -13, whil cancers -33. Resuclusions ,15, suchTP53 codon 72 polymorphism and OSCC, although subgroup analysis revealed a positive correlation between the polymorphism and HPV status in OSCC patients. Our results are in contrast with those reported by Zhou et al. [A total of 2,298 OSCC patients and 2,111 controls were included in our meta-analysis. Results of the overall population demonstrated a negative association of the u et al. , which wu et al. . Moreoveu et al. ,15, our TP53 codon 72 polymorphism was associated with HPV-related OSCC susceptibility cases. However, because there is no association between this polymorphism and non-HPV OSCC cases, it is currently unclear whether the polymorphism is merely a marker of HPV-related OSCC. Further research is warranted to investigate this relationship.The relationship between HPV and OSCC has been previously established . Our metTP53 aberration in 26 OSCC cases [TP53 contains multiple polymorphisms in addition to the one in codon 72.In 2011, Heah et al. found a significant correlation between p53 expression and CC cases . This fiOur meta-analysis has a number of limitations. First, like all meta-analyses, it is a secondary retrospective study that is limited by various factors including quality of the original studies, study population differences, and the measurement tools used. Second, statistical heterogeneity is substantial, although this is extremely common in meta-analyses of genetic association studies. We therefore performed subgroup analyses to consider the factors that may have contributed to the high degree of heterogeneity. Third, our included studies lacked comprehensive genotype information so the results of our meta-analysis were analyzed using unadjusted data; hence, we could not generate a more accurate analysis based on other adjusted factors. Finally, the sample size of our meta-analysis is relatively small and studies published in languages other than Chinese and English were not considered for inclusion.TP53 codon 72 polymorphism and OSCC susceptibility in Asians, although subgroup analysis demonstrated an association between the polymorphism and HPV-related OSCC patients. Because of the numerous limitations of this meta-analysis including small sample size and substantial statistical heterogeneity, our results should be interpreted with caution and further data from high-quality, well-conducted clinical studies of adequate statistical power are needed.Our meta-analysis showed a lack of association between the Arg: Arginine; CI: Confidence interval; HPV: Human papillomavirus; HWE: Hardy-Weinberg Equilibrium; OR: Odds ratio; OSCC: Oral squamous cell carcinoma; PCR: Polymerase chain reaction; PRISMA: Preferred Reporting Items for Systematic Reviews and Meta-Analyses; Pro: Proline; RFLP: Restriction fragment length polymorphism; SSCP: Single strand conformation polymorphism; TP53: Tumor protein p53.The authors declare that they have no competing interests.XTZ and YMN extracted the data and wrote the manuscript. WL and PLG performed statistical analysis. XTZ and WL carried out the systematic literature search and data collection. YG and WDL reviewed the manuscript. All authors approved the final manuscript.The pre-publication history for this paper can be accessed here:http://www.biomedcentral.com/1471-2407/14/469/prepub


Although several single-nucleotide polymorphisms in microRNA (miRNA) genes have been associated with primary hepatocellular carcinoma, published findings regarding this relationship are inconsistent and inconclusive.The high-resolution melting (HRM) analysis was used to determine whether the occurrence of the SNPs of miR-146a C > G (rs2910164), miR-196a2 C > T (rs11614913), miR-301b A > G (rs384262), and miR-499 C > T (rs3746444) differs in frequency-matched 314 HCC patients and 407 controls by age and sex.−6). Moreover, the allele frequency of miR-499 C > T was significantly different between the two groups, and the HCC risk of carriers of the C allele was higher than that of carriers of the T allele . Further, we found that the activated partial thromboplastin time (APTT) in HCC patients with miR-196a2 CC genotype was longer than patients with TT genotypes (P < 0.05), and HCC patients with miR-499 C allele had higher serum levels of direct bilirubin, globulin, γ-glutamyltranspeptidase, alkaline phosphatase, and lower serum cholinesterase (P < 0.05).The groups’ genotype distributions of miR-196a2 C > T and miR-499 C > T differed significantly (P < 0.01), both of them increased the risk of HCC in different dominant genetic models (P < 0.01); compared with individuals carrying one or neither of the unfavorable genotypes, individuals carrying both unfavorable genotypes (CT + CC) had a 3.11-fold higher HCC risk , 1.89–5.09; P = 7.18 × 10Our findings suggest that the SNPs in miR-196a2 C > T and miR-499 C > T confer HCC risk and that affect the clinical laboratory characteristics of HCC patients.The online version of this article (doi:10.1186/1471-2407-14-643) contains supplementary material, which is available to authorized users. Hepatocellular carcinoma (HCC) is the third most common cause of cancer-related mortality worldwide . In the et al. also found the significant association between the SNP in miR-196a2 and increased susceptibility to colorectal cancer and HCC [Previous studies have shown that single-nucleotide polymorphisms (SNPs) in microRNAs (miRNAs) may contribute to tumorigenesis owing to their ability to change the expression, regulation, and/or function of miRNAs –9. miRNA and HCC . miR-146 and HCC , 30. The and HCC , 26. miR and HCC  and was  and HCC .However, the meta-analysis revealed that the miR-146a C > G (rs2910164) variant was associated with a decreased HCC risk among Asian and male populations and no significant association was observed between the SNP and risk for HCC in the female populations . They haAlthough previous studies analyzed the relationship between different miR-499, miR-196a2 and miR-146a genotypes in different patient populations, their findings were inconsistent, and they did not investigate whether the genotypes affected patients’ clinical characteristics. To determine the role of miRNA SNPs in HCC, we performed a case–control study in which we used a high-resolution melting (HRM) genotyping method to investigate the relationship between the SNPs of four miRNAs  . Informed consent was obtained from all participants at interview, as well as at time of biospecimen collection. We included 314 patients who were diagnosed with HCC at Zhongnan Hospital between 2005 and 2012. All patients had pathologically confirmed HCC and underwent liver resection. The American Joint Committee on Cancer’s TNM  staging system and the Barcelona Clinic Liver Cancer (BCLC) staging system were used to stage patients’ HCC. From these patients we collected 314 formalin-fixed, paraffin-embedded (FFPE) tissue samples . The control group consisted of 407 participants randomly selected from healthy individuals enrolled in an HCC screening program who had no history of cancer or chronic disease. Ethylenediaminetetraacetic acid–anticoagulated peripheral blood samples were collected from the control group. Additionally, we collected 39 tumor tissue samples and peripheral blood samples from the same HCC patients. All participants’ hepatitis B surface antigen/hepatitis B virus statuses were assessed by a chemiluminescent enzyme immunoassay. The available preoperative biometrical characteristics and clinical data of the HCC patients and controls are shown in Additional file We used commercially available DNA extraction kits to extract genomic DNA from FFPE tissue samples  or peripheral blood samples  according to the manufacturer’s instructions. We used a DU 530 spectrophotometer  to quantify the concentration of DNA; absorbance readings of the DNA extracts at 260 nm indicated that the DNA concentration was about 720.94 μg/mL. The extracted DNA samples were frozen at −20°C without repeated freeze-thawing cycles until subjected to assay.2+, 2 μL of 5× polymerase chain reaction buffer, 1.0 U of polymerase enzyme, and 1× LCGreen Plus + dye (Idaho Technology). Polymerase chain reaction cycling included an initial denaturation at 95°C for 2 min followed by 45 cycles of 15 seconds at 95°C, 15 seconds at the respective annealing temperatures  in tumor and blood samples from HCC patients. We initially tested the concordance between genotypes from 39 paired tumor and blood samples using the k statistic. Then, we investigated the four miRNAS’ SNPs in the population of 314 HCC patients with FFPE samples, and 407 controls with peripheral blood samples. The primers used for the HRM analysis are shown in Additional file For quality control, DNA samples with different known genotypes were included as internal standards in each experiment. A duplicate control without a DNA template was also included in each run to test for contamination and to assess the formation of any primer dimer.http://bioinfo.iconcologia.net/snpstats/start.htm), to analyze the four miRNAs’ SNPs. All statistical tests were two-sided, and P values of less than 0.05 or Bonferroni correction–adjusted P values of less than 0.05 were considered statistically significant.We used the statistical software program SPSS 17.0 for Windows to perform all statistical analyses . Differences in the clinical characteristics and genotypes between the HCC patients and control participants were evaluated using the Student t-test or one-way ANOVA (for continuous variables) and Pearson chi-square test . The Pearson chi-square test was also used to determine whether the allele frequencies in the control group were in Hardy-Weinberg equilibrium (HWE). We used logistic regression analysis with adjustment for possible confounders (sex and age) to determine whether the genotypes of the four SNPs were associated with HCC risk; the results are presented as odds ratios (ORs) and 95% confidence intervals (CIs). To compare the clinical characteristics of HCC patients who had different genotypes, we performed a K-independent non-parametric analysis for skewed distribution. We also used SNPStats, a Web-based SNP analysis software program , which suggests that these SNPs are correlated with HCC risk.After adjustment for confounding factors (sex and age), the results of the risk estimation analysis based on genotype distribution, allele frequency, and genetic model by logistic regression analysis are shown in Table −4) and that of individuals with either CT or CC genotype in dominant model . We also found that the HCC risk of individuals with CT genotype was significantly higher than that of individuals with either CC or TT genotype in overdominant model .For the miR-196a2 C > T (rs11614913) polymorphism, the HCC risk of individuals with TT genotype was significantly lower than that of individuals with CT genotype in codominant model  in dominant model. We also found that the HCC risk of individuals with either CC or TT genotype was significantly lower than that of individuals with CT genotype in overdominant model . Additionally, the minor C allele of miR-499 (rs3746444) was associated with a higher risk of HCC .For the miR-499 C > T (rs3746444) polymorphism, the HCC risk of individuals with TT genotype was significantly lower than that of individuals with CT genotype in codominant model  and that of individuals with either CT or CC genotype  polymorphism, the HCC risk of individuals with CG genotype was significantly lower than that of individuals with CC genotype in codominant model  and that of individuals with either CG or GG genotype in dominant model . Moreover, the HCC risk of individuals with CG genotype was significantly lower than that of individuals with either CC or GG genotype in overdominant model .In addition, the results of a logistic regression analysis were consistent with those of the SNPStats analysis .−6)  had significantly different clinical characteristics  and hepatitis B virus–negative (n = 49). We found no significant difference in the genotype distributions of the four SNPs between hepatitis B virus–positive and hepatitis B virus–negative HCC patients (P > 0.05). We also found no significant association between the TNM or BCLC tumor stage and the HCC risk of patients with different genotypes (P > 0.05).Because the findings of previous studies regarding the roles of miRNA SNPs in HCC were inconclusive or inconsistent, they seem to be one of the underpinnings of the rationale for guiding us in the present study. We used HRM methods to detect the SNPs of miR-196a2 C > T, miR-499C > T, miR-146a C > G, and miR-301b A > G in HCC. HRM has been developed for the detection of DNA sequence variants and it was applied first for genotyping in 2003 , which iOur study demonstrates that miR-196a2 C > T and miR-499 C > T increase HCC risk. The HCC risks of participants who had the variant heterozygous CT genotype of miR-196a2 or miR-499 were significantly higher than those of participants who had the wild-type homozygous TT genotype of miR-196a2  or miR-499 . These results are in agreement with those reported for Chinese HCC patients. In male Chinese patients with HBV infection, the risk of HCC was significantly higher in patients with the CC genotype or carrying C allele for miR-196a2 than those with the TT genotype or T allele . SimilarOur results also suggested that the HCC risks of participants with CG genotype of miR-146a were lower than those with TT genotype . A recent meta-analysis indicated a similar result that the miR-146a C > G variant was associated with a decreased HCC risk among Asian populations . HoweverIn the present study, the distributions of the miR-301b genotypes in HCC patients and control participants did not differ significantly. The SNPs of miR-196a2 C > T, miR-499 C > T, and miR146a C > G are all located in 3p mature miRNA regions and may influence both the binding of target mRNAs to 3p and the pre-miRNA maturation of 5p and 3p. However, the SNP of miR-301b A > G is located in the miRNA flanking region. This may explain the lack of a significant difference in the distributions of the miR-301b genotypes between the two groups; perhaps this SNP did not change the maturation of the miR-301b and thus did not influence the binding of target mRNAs to 3p.In addition, the clinical characteristics of patients with different miRNA genotypes were different, and these characteristics were correlated with different genotypes. The patients with CC genotype of the miR-196a2 C > T (rs11614913) had significantly longer APTT. For the miR-499 C > T rs3746444), we also demonstrated that the patients with TT genotype had lower direct bilirubin, globulin, γ-glutamyltranspeptidase, alkaline phosphatase, and higher cholinesterase. We firstly verified the differences in coagulation function and liver function parameters between patients with TT genotype of the miR-196a2 C > T (rs11614913) or miR-499 C > T (rs3746444) and the patients with either CT or CC genotype differed significantly. The increased total bilirubin (P < 0.0001) and decreased albumin (P < 0.0001) were related to poor prognosis in patients with HCC . On the 4, we alsOur findings suggest that miR-196a2C > T and miR-499C > T increased HCC risk, and different genotypes of the SNPs in three miRNAs affected the clinical laboratory characteristics of HCC patients. It is the first study to demonstrate the relationship between different genotypes and the clinical laboratory characteristics of HCC patients. Future studies should identify the specific mechanism underlying miR-196a2C > T and miR-499 C > T genotypes as well as altered clinical laboratory characteristics, which should provide valuable information facilitating the early detection and diagnosis of HCC.It is well known that cancer tissues show frequent mutations even at SNP sites and the sequence variations in tumor tissues maybe be different from those of normal blood samples, which will almost certainly lead to questions of how to justify the tissue-with-blood comparisons. However, in this study, we compared the reliability of genetic studies done on biobanks comprised of FFPE autopsy tissue with banks of blood samples from the same donors, and investigated the association of four miRNAs’ SNPs with HCC risk. Our data suggested that the genotypes of miRNA’s SNPs were almost identical in HCC tissue and peripheral blood samples from the same patients (n = 39). The similar results were performed by Sjöholm et al., which showed that DNA from all plasma  and serum  samples gave identical genotyping results as obtained from tissue DNA from the same subject by comparison of archival plasma and FFPE tissue for genotyping in HCC, who also reported 100% each-way matching .Finally, our results were based on a small sample size. Further validation of these findings is warranted in larger studies. We will collect more FFPE tissue and blood samples from HCC patients to further address the clinical utility of the miRNA SNPs for the risk prediction of HCC.Table S2. Characteristics of hepatocellular carcinoma (HCC) patients and controls; Table S3. Primers used for high-resolution melting (HRM) analysis; Table S4. Primers used for DNA sequencing; Table S5. Comparison of the four miRNAs’ SNPs in paired tumor and blood samples; Table S6. Comparative analysis of the clinical characteristics of hepatocellular carcinoma patients with different miR-146a genotypes. (DOC 218 KB)Additional file 1: Table S1: Characteristics of the four single-nucleotide polymorphisms (SNPs);


Background. Previous studies have shown conflicting results on the association between toll-like receptor 4 (TLR4) Asp299Gly (rs4986790) polymorphism and coronary artery disease (CAD). The aim of this study was to evaluate the influence of TLR4 Asp299Gly polymorphism on CAD risk, CRP level and the number of stenotic coronary arteries, as well as to investigate whether G allele carriers would benefit more from statin treatment.Methods. PubMed, EMBASE, and CNKI databases were searched until May 2015. All the statistical tests were performed using R version 3.1.2. Odds ratio (OR) and 95% confidence interval (CI) were used to assess the association between TLR4 Asp299Gly polymorphism and CAD risk, the number of stenotic vessels, and the incidence of cardiovascular events according to statin-treated patients. Weighted mean difference (WMD) was calculated for the association between Asp299Gly and CRP level.Results. Overall, 12 case-control studies with 10,258 cases and 5,891 controls were included, and no association of TLR4Asp299Gly polymorphism with CAD was found . Also, no association was noted between Asp299Gly and CRP level . Furthermore, no synergistic effect of statin and 299Gly was reported .Discussion. This meta-analysis suggests no association of TLR4 Asp299Gly polymorphism with CAD and CRP level. It is further indicated that the G allele carriers may not benefit more from statin treatment. Further studies should include large sample size and high-quality literature to understand this issue in depth. Coronary artery disease (CAD), resulting from atherosclerosis (AS), has become the leading cause of disability and death globally . EvidencHuman TLR4 is located in 9q32-q33 region, and contains three exons. TLR4 activity and function seem to be modulated by genetic variations, especially single nucleotide polymorphisms (SNPs) . Asp299GThis meta-analysis followed the Preferred Reporting Items for Systematic Reviews and Meta-analyses (PRISMA) statement . PubMed,The inclusion criteria for the studies about the association between TLR4 Asp299Gly polymorphism and CAD were as follows: (a) published case–control studies, and the control group should be the population without CAD; (b) clear diagnosis criteria of CAD; (c) studies supplied the number of individual genotypes in CAD cases and controls; (d) sufficient data for estimating an odds ratio (OR) or weighted mean difference (WMD) with 95% confidence interval (CI); (e) written in English or Chinese. The inclusion criteria for the studies about the association of TLR4 Asp299Gly polymorphism with CRP level, the number of stenosis coronary arteries, and the incidence of cardiovascular events with statin treatment were as follows: (a) integrated data; (b) sufficient data for estimating an odds ratio (OR) or weighted mean difference (WMD) with 95% confidence interval (CI); (c) measurement data had definite unit; (d) written in English or Chinese.All reviews, case reports, animal studies and reports with incomplete data were excluded.P < 0.05 was considered as a significant deviation from HWE). Quality assessment of studies included in the meta-analysis was conducted by two authors using the Newcastle–Ottawa Scale (NOS)  was also recorded (le (NOS) . Scores This meta-analysis used four gene models: allelic model , dominant model (AA vs. GG + AG), recessive model (GG vs. AG + AA) and super-dominant model (AG vs. AA+ GG), to explore the association between the CAD risk and TLR4 Asp299Gly polymorphism. In addition, OR and 95% CI were calculated to assess the association between Asp299Gly and CAD risk, the number of stenotic vessels, and the incidence of cardiovascular events according to statin treatment; WMD was calculated for the association between Asp299Gly and CRP level.I2 statistic (P < 0.10 and I2 > 50% indicated evidence of heterogeneity)   . If no hgeneity) . Subgrouon bias) . At lastAccording to the PRISMA-statement flow diagram , 17 fullQP = 0.0019, I2 = 61.34%; dominant model (AA vs. GG + AG): QP = 0.0021, I2 = 61.13%; recessive model (GG vs. AG + AA): QP = 0.97, I2 = 0.00%; super-dominant model (AG vs. AA + GG): QP = 0.0039, I2 = 58.70%. Fixed-effects model was used for recessive model, and random-effects model was used for the other three models. G allele vs. A allele: OR =0.97, 95% CI [0.81–1.17], P = 0.75; AA vs. GG + AG: OR =0.97, 95% CI [0.80–1.18], P = 0.76; GG vs. AG + AA: OR =1.08, 95% CI [0.57–2.02], P = 0.82; AG vs. AA + GG: OR =1.03, 95% CI [0.85–1.25], P = 0.74. Overall, no association of TLR4 Asp299Gly polymorphism with CAD was observed. The results are shown in The heterogeneity of each gene model was as follows: allelic model : I2, which meant part of the source of heterogeneity was explored. However, in the dominant model and the super-dominant model, the frequency of genotypes in the case group could explain 56.86% and 63.65% of I2 respectively.The heterogeneity could not be effectively removed after subgroups were divided based on the different types of CAD. No association of Asp299Gly with CAD type was noted. To explore the source of heterogeneity further, meta-regression was used. It was found that ethnicity, region, disease category, sex ratio , and sample size ratio (cases/controls) were not the source of heterogeneity in all the models. In the allelic model, the frequency of A/G allele in the control group could explain 56.35% of One study was omitted at a time, and each of them made no obvious difference in the overall meta-analysis estimation, indicating that the results had a favorable stability.P = 0.25 for G allele vs. A allele, P = 0.25 for AA vs. GG + AG, P = 0.37 for GG vs. AG + AA, P = 0.16 for AG vs. AA + GG. Funnel plots with good symmetry and P value > 0.05 indicated no publication bias in the study.Funnel plots  intuitivAccording to the order of published years, cumulative meta-analysis was performed, and it was found that OR value and 95% CI tended to be stable, and 95% CI gradually narrowed, as shown in No association was found between the genotype frequency and the number of coronary arteries with >50% stenosis. Details are shown in Six reports  were inc article , median QP < 0.0001, I2 = 96.20%, random-effects model was used. WMD = −0.10, 95% CI [−0.62–0.41]. P = 0.69. Egger test: P = 0.77.Test for heterogeneity: Also, no significant difference was found between Asp299Gly and CRP level. Forest plot and funnel plot are shown in No synergistic effect of statin and 299Gly was found. Details are shown in The worldwide distribution of G allele is closely related to the ethnic groups, which might be the result of differences in environmental pressure during human migration . TLR4 AsIn addition, CRP, a nonspecific acute phase protein, is a useful biomarker of inflammation. It is also considered as an independent risk factor to predict first and recurrent cardiovascular events such as MI or stroke . FurtherThis meta-analysis has certain limitations as follows: first, CAD is influenced by multiple genetic mutations , so someIn conclusion, this meta-analysis suggested no association of TLR4 Asp299Gly polymorphism with CAD and CRP level. Moreover, the findings indicated that the G allele carriers might not benefit more from statin treatment. However, these results should be confirmed by conducting more high-quality studies in future.10.7717/peerj.1412/supp-1Article S1Click here for additional data file.10.7717/peerj.1412/supp-2Article S2Click here for additional data file.10.7717/peerj.1412/supp-3Supplemental Information 1Click here for additional data file.10.7717/peerj.1412/supp-4Supplemental Information 2Click here for additional data file.