Folate may involve in various aspects of carcinogenesis. However, the relationship between folate intake and risk of many cancers, including endometrial cancer, is still inconclusive. We conducted a meta-analysis to systematically review the association.Relevant studies were searched through three electronic databases  up to April 4, 2016. Population based prospective or case-control studies involving in investigating folate intake and risk of endometrial cancer were considered as eligible. Three investigators independently extracted data. Controversies were reconciled by discussing with a fourth investigator. Effect sizes of studies were pooled via a random effects model. Thereafter to explore the origin of heterogeneity among results of studies, a mixed effects model was employed with study design and dose of folate intake taken as covariates.Nine case-control studies and five cohort studies were included in the current meta-analysis. The result pooled from the highest category suggested a marginal negative association between folate intake and risk of endometrial cancer (OR=0.89 95% CI: 0.76-1.05). Based on the mixed effects model, in the highest category, the risk showed an increasing trend along with increment of folate intake .A marginally negative association was observed between folate intake and endometrial cancer, which might subject to a threshold effect. More finely designed perspective studies or randomized trials are still needed to confirm the association. As a cofactor of the de novo synthesis of purine and thymidylate and the main supplier of one carbon unite, folate plays a critical role in keeping genetic and epigenetic stability of DNA . The defAnimal experiments and epidemiological studies had shown a link between increased risk of various cancer and folate deficiency –5. HowevSome epidemiological studies suggested another story about the relationship of folate intake and cancer risk. A nested case-control study conducted in Sweden based on 226 cases and 437 matched controls suggested that the relationship between plasma folate concentration and the risk of colorectal cancer was in a bell-shaped manner . Some otBeing one of the most common gynecologic malignancy tumor, the endometrial cancer develops in over 200,000 women worldwide, deprives above 42,000 lives every year . The relThe search process was presented in Figure All studies identified were human based epidemiological studies focusing on folate intake and risk of endometrial cancer, with participants of post-menopause women. Five of 14 studies were perspective cohort studies. All five cohort studies provided dietary data from follow-up. The other nine were retrospective case-control studies, with four based on hospital and five based on population. The years of follow-up for cohort studies and duration of data collection after diagnosis for case-control studies were listed in Table Potential confounding factors including age, total energy intake, alcohol intake, and BMI were controlled or matched while estimating ORs or RRs. Twelve out of 14 studies each had a sample size of more than 200 cases , 18-25. I2=59%, Pheterogeneity=0.003). The overall result suggested a marginal negative relationship between folate intake and the risk of endometrial cancer . Meanwhile comparing to cohort studies, the mean lnOR of case-control studies was significantly lower .We also applied the mixed effects model to the second highest category. No significant publication bias was observed . The pattern of results was similar as that in the highest category, with a comparable increasing trend  and a more substantial difference of mean lnORs between different study designs .I2=30.15%, Pheterogeneity=0.15), additional covariates were not necessary. Interestingly, if we incorporated the two covariates into the mixed effects model in the same way as the two higher categories, the trend of risk relation to intake of folate turned neutral ; yet the effect of study design remained relative constant, with an estimation of -0.21 and a P value of 0.046 for the difference of mean lnOR between two subgroups.When it comes to the third higher category, The P values for Begg's test and Egger's test were 0.638 and 0.22 respectively, suggested no significant publication bias. Because of the insignificant heterogeneity (BeggP=0.67), EggerP=0.37, Figure Egger's test and Begg's test was performed in all 14 studies. Funnel plots were used for graphical representation. Significant publication bias was not observed, as suggested by Begg's test and Egger's test , EMbase (http://www.embase.com), and Web of Science (http://www.webofknowledge.com). The references of studies and relevant reviews were also checked.The process of the meta-analysis in this article was complied with the Preferred Reporting Items for Systematic Reviews and Meta-Analysis Statement (PRISMA) issued in 2009 . We systThe key words for searching were as following: (“folate” OR “folic” OR “folacin”) OR (“pteroyl-l-glutamic acid” OR “pteroyl-l-glutamate” OR “pteroylmonoglutamic acid”)) paired with  OR  AND ).The research was a human based case-control study, prospective study or clinical trial.The interests of article included investigating the association between folate intake and risk of endometrial cancer.The effect size  and their confidence intervals (CIs) were available.The inclusion criteria for studies are as following:Studies based on cell lines or animal.The article is not written with English.Reviews, case-report, editorials or commentaries.The exclusion criteria were as following:Three investigators  identified eligible articles independently according to the aforementioned criteria. Controversies on article inclusion among the three reviewers were resolved through discussing with a fourth investigator (Ying Gao).Data on the study design, country, sample size, age, food frequency questionnaire items, time frame, exclusion of hysterectomy together with folate intake level of each category, effect size , and 95% CI were collected. If crude OR or RR was reported together with multivariable adjusted-effect estimates, we chose the estimation that had been fully adjusted for the potential confounders.The quality of studies was accessed by sample size and whether important confounders such as alcohol intake, total energy intake, body mass index and age had been adjustment for.Q statistic and 2I statistic. As long as the P value for heterogeneity was smaller than 0.05 or the 2I statistic was greater than 25%, results of studies were considered as significantly heterogeneous and a DerSimonian and Laird random effects model was used to pool the results [All statistical analyses were performed with R, version 3.1.2 (Package metafor). The primary outcome was defined as odds ratio (OR). Thus risk ratios (RR) together with their variance were converted to lnORs for further analysis . Heterog results , 37. OthTo explore the origin of heterogeneity among studies, a mixed effects model was applied to three higher categories as most of candidate studies were in quartile. Since the mandatory folic fortification in North America had led to a substantial difference in folate intake of populations in different country, the dose of folate intake, as well as study design were chosen as covariates of model , 39. WitIf the population on which a study based was divided into less than four categories, for instance, in tertiles, the two lnORs were allocated for pooling results of the highest and the second highest category. At the second highest category, study of Randi et al  was exclThe results of every study in all the models above were negative variance weighted. Publication bias was detected by Egger's test and Begg's test, then graphically represented by funnel plots.


Published data on the association between Toll-like receptor 4 (TLR4) Asp299Gly polymorphism and coronary heart disease (CHD) susceptibility are inconclusive. To derive a more precise estimation of the relationship, a meta-analysis was performed. English-language studies were identified by searching PubMed and Embase databases (up to November 2016). All epidemiological studies were regarding Caucasians because no TLR4 Asp/Gly and Gly/Gly genotypes have been detected in Asians. A total of 20 case-control studies involving 14,416 cases and 10,764 controls were included in the meta-analysis. Overall, no significant associations were found between TLR4 Asp299Gly polymorphism and CHD susceptibility in the dominant model  pooled in the meta-analysis. In the subgroup analysis by CHD, non-significant associations were found in cases compared to controls. When stratified by control source, no significantly decreased risk was found in the additive model or dominant model. The present meta-analysis suggests that the TLR4 Asp299Gly polymorphism was not associated with decreased CHD risk in Caucasians. Epidemiological studies have strongly supported a pivotal role for inflammatory, innate immune, and adaptive immune mechanisms in the pathogenesis of atherosclerosis ,2. AlthoVariants in the gene encoding TLR4 may affect the development of atherosclerosis accompanied by an impaired signal transduction, but the responsible polymorphisms remain inconclusive ,7. A sinThis study was carried out and reported in agreement with the PRISMA guidelines for systematic reviews and meta-analyses . Each stWe independently evaluated eligible articles on the basis of the following inclusion criteria: 1) evaluation of TLR4 Asp299Gly polymorphism and CHD susceptibility, 2) case-control studies, and 3) sufficient published data for estimating an odds ratio (OR) with 95% confidence interval (CI).Two authors (B.W. Wu and J. Zhu) independently extracted data from all eligible studies fulfilling the inclusion criteria. Disagreement was resolved by discussion between the two authors. If these two authors could not reach a consensus, another author (H.M. Shi) was consulted and a final decision was made by the majority of the votes. Data extraction included the first author’s surname, publication year, origin region, matching criteria, demographic data, and genotyping method. For data not provided in the main text, the required information was obtained in part from supplementary online appendixes.http://www.cochrane.org/cochrane/hbook/htm).Cochrane collaboration meta-analysis review methodology was used for this study . ORs witA total of 324 unique citations were identified by our search strategy. After the initial screening, 56 potentially relevant articles were selected for further review. Among these, 36 articles were excluded according to the inclusion criteria. Overall, 20 case-control studies involving 14,416 cases and 10,764 controls were included in the meta-analysis .2=70.9%; P<0.00001). In the subgroup analysis by CHD, non-significant associations were found. When stratified by control source, no significantly decreased risk was found in dominant model.vs Asp/Asp) from 13 case-control studies . As shown in vs Asp/Asp) in patients with CHD compared with controls . In the subgroup analysis by CHD and control source, non-significant associations were found in additive model.As depicted in A single study involved in the meta-analysis was deleted each time to reflect the influence of the individual data-set to the pooled ORs, and the corresponding ORs were not significantly altered, indicating that our results were statistically robust .Begg’s funnel plot and Egger’s test were performed to access the publication bias of the studies. The funnel-plot analysis indicated that no significant publication bias was detected to influence the results of this meta-analysis.Atherosclerosis, although closely related to lifestyle and environmental factors, is also influenced by the complex patterns of inheritance ,34. In tThe present study found non-significant associations between TLR4 Asp299Gly polymorphism and CHD susceptibility in 20 case-control studies. Considering the complex nature of atherosclerosis, it suggests that TLR4 Asp299Gly polymorphism has only a minor impact on the pathogenesis of CHD.Compared with a previous meta-analysis , we inclIn conclusion, the present meta-analysis suggested that TLR4 Asp299Gly genetic polymorphism is not involved in the pathogenesis of CHD in Caucasians. However, despite the negative results, a causal relationship may exist in the development of atherosclerosis. Well-designed studies with larger samples should be conducted to confirm the results. Moreover, further studies estimating the effect of gene-gene and gene-environment interactions may eventually provide a better and more comprehensive understanding of the association between TLR4 Asp299Gly polymorphism and CHD susceptibility.


A total of 507 ESCC cases and 1,496 controls were enrolled. We found that LEPR rs6588147 AA genotype was associated with ESCC risk . In the stratified analyses, LEPR rs6588147 G>A polymorphism increased the risk of ESCC . However, LEP rs7799039 A>G decreased the risk of ESCC . In addition, LEPR rs1137101 G>A polymorphism decreased ESCC risk in some subgroups . Our findings suggest that LEPR rs6588147 G>A polymorphism is associated with the increased risk of ESCC; however, LEP rs7799039 A>G and LEPR rs1137101 G>A polymorphisms may be protective factors for ESCC.Single nucleotide polymorphisms (SNPs) in energy metabolism related gene may be key agents in the development of human malignancies. In this study, we aimed to examine the association of  In China, esophageal cancer (EC) is the fourth most commonly diagnosed cancer in males and the fifth in females, with an estimated 477,900 new patients and 375,000 related deaths occurring in 2015 [transcription factor 7-like 2 (TCF7L2) gene maps to the long arm of chromosome 10q25.3. TCF7L2 belongs to the high mobility group-box (HMGB) family [et al. reported that TCF7L2 expression was associated with a poor prognosis of ESCC [TCF7L2 rs7903146 locus might exert its enhancer function by interacting with HMGB1 [TCF7L2 single-nucleotide polymorphisms (SNPs) are proposed susceptibility factors for the development of cancer. Previous studies indicated that TCF7L2 rs7903146 (C/T) polymorphism might influence the risk of breast cancer [TCF7L2 rs290481 T>C polymorphism located on near the 3’ end of this gene. Ling et al. reported that this SNP was associated with hepatocellular carcinoma susceptibility with marginal significance [TCF7L2 SNPs and ESCC risk was not explored.The ) family  and is a) family , 7. Ishi of ESCC . A previt cancer . TCF7L2 ificance . HoweverLeptin (LEP) gene maps to chromosome 7q31.3. LEP is secreted by white adipose tissue and has been identified to be involved in endocrinologic metabolism [LEP gene might influence the risk of cancer [LEP rs2167270 G>A was associated with the risk of esophageal adenocarcinoma [LEP polymorphisms. It is suggested that rs7799039 A>G polymorphism in the upstream region of LEP gene can affect leptin expression, possibly at the transcriptional level, thereby altering adipose secretion levels of the hormone [LEP rs2167270 G>A is a 5’-utr SNP and may play regulatory roles in translation and stability of mRNA. LEPR rs1137100 G>A, rs1137101 G>A polymorphisms are missense SNPs and may alter the structure and the function of LEPR protein. Doecke et al. found LEPR rs1137100 G>A, rs1137101 G>A polymorphisms influence the risk of esophageal adenocarcinoma in Caucasians [LEPR rs6588147 G>A polymorphism locates on the intron region of LEPR gene. Slattery et al. found that LEPR rs6588147 G>A polymorphism affected risk of colon cancer among men [LEPR rs1137100 G>A, rs1137101 G>A and rs6588147 G>A polymorphisms and ESCC risk remains unknown in Asians.The tabolism . It is ttabolism , type 2 tabolism  and pathtabolism , 17. LEPtabolism . LEP comtabolism –22. Thusf cancer –25. In aarcinoma . There aucasians . LEPR rsmong men . HoweverTCF7L2, LEP and LEPR polymorphisms with the risk of ESCC in Eastern Chinese Han populations. The TCF7L2 rs7903146 C>T, rs290481 T>C, LEP rs7799039 A>G, rs2167270 G>A and LEPR rs1137100 G>A, rs1137101 G>A and rs6588147 G>A polymorphisms were genotyped by SNPscan genotyping assays in 507 ESCC cases and 1,496 non-cancer controls.In this case-control study, we aimed to examine the potential association of P < 0.05). However, the mean ± SD of height was not significant (P > 0.05). The proportion of smoking and drinking was significantly higher in ESCC patients compared with controls (P < 0.05). Locus information of TCF7L2, LEP and LEPR polymorphisms is listed in Table TCF7L2 rs7903146C>T, rs290481 T>C, LEP rs7799039 A>G, rs2167270 G>A and LEPR rs1137100 G>A, rs1137101 G>A and rs6588147 G>A SNPs were 99.50%,99.45%, 99.50%, 99.40%, 99.50%, 99.50% and 99.50%, respectively. Minor allele frequency (MAF) in controls is listed in Table TCF7L2 rs7903146C>T, rs290481 T>C, LEP rs7799039 A>G, rs2167270 G>A and LEPR rs1137100 G>A, rs1137101 G>A and rs6588147 G>A genotypes in controls conform to Hardy-Weinberg equilibrium (HWE).There were 2,003 participants in the present case-control study including 507 ESCC patients  and 1,496 non-cancer controls . The age and sex were well matched in two groups  = 1.87, 95% confidence interval (CI) = 1.02–3.43, P = 0.042 and AA vs. GG/GA: crude OR = 1.93, 95% CI = 1.06–3.53, P = 0.031 .In addition, after a logistic regression analysis, we found that TCF7L2, LEP and LEPR) polymorphisms and focused on their susceptibility to ESCC. The association between LEPR rs6588147 G>A polymorphism and the increased risk of overall ESCC was identified. We also found that LEPR rs6588147 G>A polymorphism increased the risk of ESCC in <63 years and male subgroups. LEP rs7799039 A>G was associated with the risk of ESCC in ≥63 years and BMI ≥ 24 kg/m2 subgroups. In addition, LEPR rs1137101 G>A polymorphism decreased the risk of ESCC in ever smoking and ever drinking subgroups.The pathogenesis of ESCC was very complex. Multiple factors  may contribute to the etiology of ESCC. Understanding of the individual’s heredity background may be helpful for the prevention and treatment of ESCC. In this study, we selected energy metabolism and insulin-sensibility relative gene . When each subject was interviewed, a questionnaire was used to obtain demographic variables and risk factors. And weight and height were also measured. In this study, a BMI ≥ 24 was considered as the criteria for obesity and overweight .TCF7L2 rs7903146C>T, rs290481 T>C, LEP rs7799039 A>G, rs2167270 G>A and LEPR rs1137100 G>A, rs1137101 G>A and rs6588147 G>A genotypes were assessed by the SNPscan™ kit , which is a double ligation and multiplex fluorescence PCR [TCF7L2, LEP and LEPR polymorphisms were confirmed.Genomic DNA was carefully isolated from EDTA-anticoagulated blood of recipients by using a Promega DNA blood mini kit . ence PCR . For quaTCF7L2, LEP and LEPR genotypes, sex, age, BMI, smoking and drinking status) were compared with Chi-square test (χ2). Deviations from the HWE for TCF7L2, LEP and LEPR genotypes distribution in controls were evaluated by an internet-based calculator (http://ihg.gsf.de/cgi-bin/hw/hwa1.pl) [TCF7L2 rs7903146C>T, rs290481 T>C, LEP rs7799039 A>G, rs2167270 G>A and LEPR rs1137100 G>A, rs1137101 G>A and rs6588147 G>A polymorphisms with ESCC susceptibility were evaluated by crude ORs and 95% CIs. Multivariate linear regression adjusted for age, sex, BMI, alcohol use and smoking status was used to determine the relationships between TCF7L2 rs7903146C>T, rs290481 T>C, LEP rs7799039 A>G, rs2167270 G>A and LEPR rs1137100 G>A, rs1137101 G>A and rs6588147 G>A polymorphisms and ESCC risk with quantitative traits. Data analysis was conducted with SAS software for windows . A P < 0.05 (two-tailed) was accepted as the criterion of statistical significance.Continuous variables  are expressed as mean ±standard deviation (SD). Comparisons between ESCC patients and controls were carried out with Student’s t-test. The categorical variables (e.g. hwa1.pl) –49. The


We aimed to assess the association between allergic conditions and risk/mortality of colorectal cancer (CRC). A systematic literature search was conducted using Pubmed and Embase to identify relevant studies. Prospective studies assessing the association between allergic conditions and risk/mortality of CRC were included. Risk ratios (RRs) were pooled with either a fixed- or a random-effects model according to heterogeneity. A total of 515379 participants and 10345 CRC cases from 12 studies were included in the analysis of CRC risk, while four studies with 1484741 individuals and 30040 CRC deaths were included in the analysis of CRC mortality. The pooled RR for the association between allergic conditions and CRC risk was 0.88 (95% CI 0.83–0.92). The inverse association was observed both in colon cancer  and rectal cancer . Moreover, no gender difference was observed in the analysis of CRC risk . And allergic conditions were also found to be inversely associated with CRC mortality . In conclusion, the current meta-analysis provides further evidence that allergic conditions were inversely associated with CRC risk and mortality. A number of lifestyle and environmental risk factors have been identified for CRC, including smoking, diet, obesity, diabetes, physical activity, and stain or aspirin use4. However, the exact mechanisms of CRC remain unclear. Allergic conditions including hay fever, asthma, atopic dermatitis, food allergy and drug allergy, may indicate an enhanced immune system, which may contribute to recognize and remove malignant cells and thus reduce risk of malignancies. Several studies reported a lower incidence of all malignancies among patients with allergic conditions compared with general population6, while some other studies revealed a positive or non-significant association between allergic conditions and malignant disease8. It was suggested that the association between allergic conditions and malignancy risk may be organ specific10. For example, allergic conditions were reported to be associated with a lower risk of pancreatic cancer9 but a higher prostate cancer risk10.Colorectal cancer (CRC) is the third most common cancer in men and second most common cancer in women worldwide, with approximately 373,600 men and 320,300 women dying from CRC in 201214. Considering these uncertainties, we performed a systematic review and meta-analysis to clarify the risk/mortality of CRC in patients with allergic conditions.In recent years, a number of studies investigated the association between allergic conditions and CRC risk or mortality with inconclusive results15.This meta-analysis was planned, conducted and reported in adherence with the PRISMA statementA systematic literature search was performed in the Pubmed and Embase databases  without restrictions. The following search terms were used:  AND . Moreover, reference lists of retrieved articles and relevant reviews were also scanned to avoid missing studies. We carefully examined the retrieved articles to exclude duplicate studies. Two authors independently screened the titles and abstracts of the articles selected from the initial search for potential inclusion, and full articles were reviewed subsequently to determine inclusion of eligible studies.Criteria for inclusion were as follows: (1) the study was a prospective cohort or nested case-control study; (2) the study assessed the association between allergic conditions and risk and/or mortality of CRC; (3) relative risk (RR) or hazard ratio (HR) were reported or could be calculated. If data were reported in two or more studies, we only extracted the most detailed or recent information.16. Studies were categorized into high quality if the score was 7 points or more, as medium quality if the score was 4 to 6 points and as low quality if the score was less than 4. The data extraction and quality assessment were conducted independently by two authors, and any discrepancies were resolved by discussion with a third investigator.We extracted the following information from each study: first author, year of publication, country, age and gender of participants, number of participants, number of CRC cases or deaths, variables adjusted for in the analysis, type of allergies evaluated in the study and RR (or HR) estimates with 95% CIs for CRC risk/mortality. RRs (HRs) reflecting the greatest degree of control for potential confounders were adopted in the meta-analysis. The Newcastle-Ottawa Scale was used to evaluate the quality of the included studies2 test; p ≤ 0.05 and/or I2 > 50% were defined as significant heterogeneity17. Study-specific RR (HR) estimates for allergic conditions and CRC risk/mortality were pooled using a random-effects model if there was significant heterogeneity; otherwise, a fixed-effects model was applied. Subgroup analyses were further conducted for the association between allergic conditions and CRC risk, stratifying by gender, anatomical sites , geographic region and type of allergic conditions. Moreover, sensitivity and subgroup analyses were performed to explore the source of heterogeneity in the meta-analysis. To evaluate the publication bias risk, funnel plots and Egger’s linear regression method were conducted. Two-sided p values were calculated, with a p < 0.05 considered significant for all tests. All analyses were performed using the Stata software .The extent of heterogeneity across studies was assessed using the chi-square test and I25. Figure A literature search of Pubmed and Embase databases identified 2327 results, of which 46 were potentially eligible articles for further review. Among them, 32 papers were excluded because of the following reasons: the study was not an original study (n = 5), the study was not a prospective cohort or nested case-control study n = 14), the study did not report the association between allergic conditions and CRC risk/mortality (n = 8), or an RR (HR) estimate was not reported and could not be calculated (n = 5). Finally, a total of 14 studies met the inclusion criteria and were included in this meta-analysis4, the stTwelve of the 14 included studies evaluated CRC risk in patients with allergic conditions, while 4 investigated CRC mortality . All the studies were conducted in Western countries: eight in USA, five in European countries and one in Australia. Among the included studies, twelve were cohort studies and the other two were nested case-control studies. Two studies enrolled female participants only, whereas the other studies included both males and females. Sample size of the included studies ranged from 3308 to 199112. All the included studies investigated both colon and rectal cancer. The methodological quality of the included studies was shown in Supplementary Table 2 = 30.7%, p = 0.146)   indicated no evidence of publication bias.A total of 12 studies assessed the association between allergic conditions and risk of CRC. These studies comprised 515379 participants, of whom 10345 developed CRC. Suggested by the pooled result, individuals with allergic conditions were at a lower risk of CRC   and in the studies conducted in the USA . Sensitivity analyses were performed to explore the potential source of heterogeneity. We found that after excluding the study by Jacobs et al.14, no significant heterogeneity was found in the analysis of colon cancer , while heterogeneity in the analysis of studies from USA decreased after excluding the study by Prizment et al.19 . The results of subgroup analyses were shown in Table Subgroup analyses revealed that individuals with allergic conditions had a lower risk of both colon cancer  and rectal cancer . We also found an inverse association between allergic conditions and CRC risk both in males  and in females . With regard to specific allergic conditions, skin allergy , hay fever , and asthma  were not significantly associated with CRC risk. In the subgroup analyses, significant heterogeneity was observed in the analysis of colon cancer    without significant heterogeneity 29. Allergic conditions appear to have dual roles in CRC30. Allergies may cause chronic inflammation and stimulate cell growth, which can predispose an individual to malignant disease30. On the other hand, two hypotheses, “immunosurveillance” hypothesis and “prophylaxis” hypothesis, have been proposed to explain the inverse association between allergic conditions and cancer28. The “immunosurveillance” hypothesis proposes that people with allergic conditions have heightened immune system, which contributes to detect and eradicate pre-malignant cells and thus reduces cancer risk32. Another potential explanation, known as prophylaxis hypothesis, asserts that allergic responses in mucosal surfaces can lead to more rapid clearance of toxins, pathogens, and foreign particles before they initiate carcinogenesis28. The protective effect of allergic conditions on CRC could also be explained by the antitumor effect of type I immunoglobulin E-mediated immune activity11. Atopic allergens can be taken into the digestive system and lead to type I immunoglobulin E-mediated hypersensitivity reactions33. A previous study found that coculture of eosinophils with colorectal carcinoma cells resulted in secretion of eosinophil cationic protein and granzyme A, which exerts eosinophil tumoricidal activity toward CRC cells34. Besides, IgE can bind to cell surface IgE receptors such as CD23 and FceRI, which engage several types of effector cells against cancer35. Moreover, it is also indicated that treatment with tumor-specific mouse IgE antibody could inhibit human colorectal tumor xenograft growth36. More studies are warranted to further clarify the exact mechanisms for the antitumor effect of allergic conditions.Though a large number of studies have investigated the role of allergic conditions in cancer, the mechanisms remain unclear37. However, most of the included studies were case-control studies, while only 8 were cohort/nested case-control studies (6 were prospective and 2 were retrospective). Retrospective studies tend to have biases in control selection and recall of former exposure to risk factors38. Prospective studies can avoid these problems, thus evidence from prospective studies is generally considered to be stronger than that from retrospective studies. In the current meta-analysis, all the included studies were prospective cohort or nested case-control studies with a substantial number of participants, which increased the statistical power.A previous meta-analysis evaluated the association between atopic diseases and breast, prostate, and colorectal cancers, finding no significant association between atopic diseases and CRC risk11. Endoscopy use could reduce CRC risk by removal of adenomas, while it may also increase the rate of CRC detection. This may lead to over-estimation or under-estimation in the effect of allergic conditions. Second, the types of allergic conditions and the diagnostic methods were different across the included studies. An inverse association between any allergic condition and CRC risk/mortality was observed, while no significant association was indicated for specific allergic conditions including asthma, hay fever and skin allergy. Other kinds of allergies, such as food allergy, were not evaluated because of the limited data. An individual participant data (IPD) meta-analysis might be helpful to clarify this issue. Finally, all the included studies were conducted in Western countries, thus, the results of the current study should be interpreted with caution for population from other countries.However, there are still several limitations that should be considered when interpreting the results. First, though most of the studies included in the current meta-analysis were adjusted for other known risk factors of CRC (shown in Table In conclusion, the current meta-analysis of prospective studies suggested an inverse association between allergic conditions and risk/mortality of CRC. The protective effect of allergic conditions was observed both in colon cancer and rectal cancer. These findings further highlight a potential protective role of the reactive immune system in CRC.Supplementary File


P = 0.020). However, this association became non-significant after excluding the study with the smallest odds ratio. Subgroup analyses revealed a significant decrease in risk of lung cancer . Our study indicates that miR-423 rs6505162 might be associated with a reduced risk of cancers, however, this finding need to be evaluated further in larger samples, especially subgroup analyses. In addition, cancer-specific functional studies are especially needed to reveal the underlying mechanisms between miR-423 and the etiology of cancer.The association between miR-423 polymorphism (C > A) and the risk of different cancers are still controversial. We performed a meta-analysis to clarify its association with multiple cancer risks. PubMed and Embase  were searched. A total of 17 studies from 16 articles, consisting of 8,582 cases and 10,291 controls, were finally qualified and enrolled in this meta-analysis. The pooled results showed that the miR-423 AA genotype was associated with decreased cancer risk under the recessive model (odds ratio [OR] = 0.87, 95% confidence interval [CI]: 0.78~0.98,  Cancer poses a major threat to public health worldwide, and its burden continues to increase . The caupre-miR-423 and maps to 17q11.2, with a nucleotide alteration from C to A. The aberrant expression of both mature forms of the miR-423 (named as miR-423-3p and miR-423-5p) has been observed in numerous cancer types [The rs6505162 SNP is located in the er types –17. A grer types –33. Abouer types , 31, 33.er types , 24, 28 er types , 30, 32.er types  we perfoThe selection flow of studies was summarized in Figure All of the eligible studies were issued in English except for one issued in Chinese . A totalP = 0.121, Figure P = 0.020)  Figure .P = 0.161; OR = 0.91, 95% CI: 0.78~1.06, P = 0.231, respectively). With respect to cancer type, a significant decrease in cancer risk was found only in lung cancer patients . The results of meta-regression analysis showed that cancer types and ethnicity do not affect the association between miR-423 rs6505162 polymorphism and cancer susceptibility (both P > 0.05).Subgroup analyses were performed based on ethnicity and cancer type Table . There wP = 0.017), indicating that patients carrying miR-423 rs6505162 AA genotype may have a decreased cancer risk compared with those carrying the CC/AC genotype.Sensitivity analysis performed by excluding the “low quality” study with a NOS score of < 6 . The pooP = 0.114).We also did another sensitivity analysis by excluding the study with the smallest OR in the recessive model . The pooIn addition, 9 studies reported the adjusted ORs under dominant model, while 8 studies reported the adjusted ORs under recessive model Table . The pooP = 0.952 for the dominant model and P = 0.906 for the recessive model).Visual inspection of funnel plots and Egger's test was used to evaluate the publication bias in our meta-analysis. Taking the recessive genetic model as an example, the funnel plot is displayed in Figure To the best of our knowledge, this meta-analysis is the first to comprehensively evaluate the association between miR-423 rs6505162 (C > A) and all cancer risk. In this meta-analysis, 17 eligible case-control studies involving 8,582 cases and 10,291 controls were analyzed. Our major finding suggested that a significant association between miR-423 rs6505162 and cancer susceptibility was observed in recessive models. Subgroup analyses also linked the A allele with a significant decreased risk of lung cancer.http://osse.bii.a-star.edu.sg/calculation2.php), assuming that the minor allele frequency (MAF) of the A allele is 0.2785 and 0.2980 in cases and controls, respectively, according to our own data. The power before and after excluding the above study were just 81% and 79% in corresponding sample size, respectively. Another sensitive analysis revealed that the pooled ORs of recessive model analysis kept significant, based on the 8 studies reported the adjusted ORs. With these evidences, as well as the almost marginal significance in original recessive model, the non-significant result (P = 0.114) in our sensitive analysis may be mostly due to the insufficient sample size. Furthermore, our result was consistent with another cross phenotype meta-analysis (CPMA), which is based on an in silico re-analysis of the human genotyping data downloaded from dbGAP [About seven of 17 included studies reported significant association between rs6505162 and risk of different cancers –28, 30. om dbGAP .The meta-analysis was also performed in different populations and cancer types to further eliminate heterogeneity. When stratified by race, no significant association was detected in any genetic model involving any population, but an exception was found for both the dominant model and the recessive model in a black population (both ORs < 1). However, interpreting this result should be done cautiously, due to the fact that the only included study had a limited sample size of about 368 cases and 583 controls . During NSRP1), [miR-423 in cell lines from breast cancer and endometrial carcinoma, [but the SNP was not correlated with expression of miR-423 in esophageal squamous cell carcinoma [TIF1 (gastric cancer), [ATG7 (hepatocarcinoma), [KLF2 (ovarian cancer) [It is known that rs6505162 lies within the first intron of the gene of nuclear speckle splicing regulatory protein (NSRP1), , 38 and NSRP1), , 39. HowNSRP1),  oral canNSRP1),  inconsisNSRP1), –43 whileNSRP1),  laryngeaNSRP1),  female gNSRP1), –47 and mNSRP1), –50, 51. NSRP1),  as oncogNSRP1), , 51 but NSRP1), , 53. Addrcinoma,  but the arcinoma . Moreovearcinoma  and gastarcinoma  but it iarcinoma . The procancer),  ATG7 (hecinoma),  and KLF2We believe that this is the first quantitative assessment focused on the association between miR-423 rs6505162 alleles and all types of cancer. Our results are reliable for the following reasons. First, the genotype distributions in the controls of this SNP were all mostly consistent with HWE. Second, no apparent publication bias was observed by either Begg's funnel plot or Egger's test. Third, all included studies used high quality genotyping methods according to the methodological quality assessment. Some limitations of this study should be also acknowledged. As noted, only 16 articles with 17 studies were included, and the data in some analyses and subgroup analyses  were relatively insufficient. Almost 70% of the participants included in this meta-analysis were from one ethnical group (Chinese); therefore generalizations should be made cautiously. The lack of original data limited further analysis of the potential interactions. Finally, although the statistical tests for publication bias were not significant, publication bias may still exist due to that studies with negative results often have less chance for publication. Considering all these factors, our results should be interpreted with caution, but we believe these findings could help to explain the association between miR-423 rs6505162 and cancer risk.In conclusion, this meta-analysis indicated that miR-423 rs6505162 C > A may reduce the risk of cancer, especially for lung cancer. The high heterogeneity among cancer types indicated that this polymorphism might play different roles in different cancers. However, this finding needs to be evaluated further in larger samples, especially for subgroup analyses. In addition, cancer-specific functional characterizations are simultaneously needed to reveal the underlying mechanisms between miR-423 rs6505162 and the etiology of cancer.A comprehensive search was conducted to identify all eligible publications in PubMed and Embase electronic databases as of 10th September, 2016. The medical subject headings (MeSH) and free-text words were used. Search terms mainly included  AND . We also carefully checked references of the retrieved articles to find additional eligible studies. During the course of literature search, no language or other limits were set.The study was considered eligible if it met the following criteria: (1) designed as case-control or cohort study, (2) evaluated the association between miRNA-423 polymorphism and cancer risk, (3) provided sufficient data  for calculating the OR and its 95% CI. We excluded letters, comments, correspondence, conference reports and laboratory studies or articles that did not contain enough data with which to compute the ORs.Two investigators independently extracted the following items from each eligible study: surname of first author, country of the investigation, year of publication, ethnicity, genotyping method, genotype distribution and HWE in the control group. A discussion was carried out to achieve consensus when discrepancies were noted.Two investigators independently evaluated the quality of eligible studies using the NOS,  which waP < 0.05 was considered a significant departure from the HWE. The I2 statistic and Q test were used to measure the between-study heterogeneity. If I2 < 50% and P > 0.1, the heterogeneity was considered mild, and the summary ORs were combined under a fixed-effects model, otherwise a random-effects model were used. The Z test was used to assess the statistical significance of pooled ORs, and two-tailed P-values < 0.05 were considered significant.Crude ORs together with their corresponding 95% CIs were calculated to assess the strength of association between miRNA-423 rs6505162 and overall cancer risk under dominant and recessive models. HWE was examined for each study by the Chi-square test in the control groups, and Both subgroup analyses and meta-regression analysis were performed to explore potential sources of heterogeneity in ethnicity and cancer types. Sensitivity analyses were performed in two steps: (1) excluding the studies with “low quality”, and (2) excluding the studies with the biggest or smallest OR in genetic models with statistically significant findings. Furthermore, we also extracted the adjusted OR and 95% CIs from the original literature to evaluate the stability of the findings. Visual inspection of funnel plots and Egger's regression asymmetry test were applied to assess potential publication bias. STATA 14.0  was used for statistical analyses.


DNA ploidy, a DNA flow cytometry parameter, reflects tumor cell cycle. In breast cancer (BC), ploidy status characterizes genotypic stability and potential metastatic capacity. It is suggested that aneuploidy is an independent prognosticator for BC patients and could aid for individualized medicine. There are extensive studies concerning the prognostic significance of DNA aneuploidy, however, its clinical utility remains controversial. Herein we conducted a meta-analysis to determine the correlation between DNA ploidy status and BC characteristics and survival.The electronic databases PubMed, EMBASE, and Web of Science were searched for relevant studies. The major investigated parameters were the BC aneuploidy rates in relation to tumor stage, size, lymph node metastasis, grading, estrogen receptor (ER) status, disease-free survival (DFS), and overall survival (OS). Hazard ratios (HRs) and the corresponding 95% confidence intervals (CIs) for DFS and OS were extracted from each study before meta-analyzed. Risk ratios (RRs) were computed using the fixed-effect or random-effects model according to data heterogeneity, and the Mantel-Haenszel or the inverse-variance method was adopted where appropriate to obtain pooled estimates using RevMan 5.3. The Egger's test was conducted with Stata 11.vs. stages II-IV, RR=0.84; 95% CI, 0.74 to 0.96; P=0.01), larger tumors , lymph node metastasis , poorer tumor proliferation , and ER− status  were significantly more frequently aneuploid. BC patients with diploid tumors had better clinical outcomes than those with aneuploid cancers. The pooled HR estimates were0.73 (P<0.0001) for DFS and 0.72 (P=0.0001) for OS, respectively.Pooled analyses of data from 29 studies involving a total of 141,163 cases showed that BC patients with more advanced tumors (stage I This meta-analysis implies that DNA aneuploidy is a significant predictor for BC progression and survival, and should be focused on in the therapeutic planning. Breast cancer (BC) is one of the most common malignancies and a leading cause of cancer-related mortality among women worldwide , 2. TumoDNA cytometric techniques, such as flow cytometry (FCM) and image DNA cytometry, are generally used in tumor cytology . ClinicaAs DNA cytometry is widely used in tumor pathology, DNA ploidy emerges as a prominent marker to reflect tumor proliferation. DNA ploidy is the set number of cell chromosomes. Due to the fact that DNA ploidy reflects the cell cycle of a tumor, it is supposed to reflect the biological behavior of the malignancy. DNA ploidy status is mostly defined as follows: the determination of DNA ploidy refers to DNA index (DI), which represents the ratio of the DNA content of G0/1 tumor cells to the reference G0/1 diploid peak (2n) for each specimen as shown in flow cytometry plots . The result from the Egger's test showed that there was no indication of a bias (P = 0.956).Six studies , 34, 35 P < 0.00001, random-effects model; Figure P = 0.013; RD = −0.14; 95% CI, −0.15 to −0.13; P < 0.00001, fixed-effect model). The Egger's test showed that there was no significant bias (P = 0.408).Eleven studies , 33, 34 P < 0.00001; sign test: P = 0.074; Figure P < 0.00001), based on a fixed-effect model.There are 8 studies , 25, 33 P < 0.00001, fixed-effect model; Figure P = 0.074; RD = 0.21; 95% CI, 0.13 to 0.28; P < 0.00001, random-effects model). Besides, aneuploidy was significantly more frequent in G3 tumors than in G1 tumors . Similar results were observed when comparing G3 and G2 tumors  based on a random-effects model. The results from the Egger's tests showed that there was no indication of biases for these two meta-analyses .Eight studies , 33–35 p+ and ER− BCs. The pooled analysis uncovered that aneuploidy was significantly more frequent in ER− tumors than in ER+ tumors  based on a random-effects model. The Egger's test demonstrated no significant bias (P = 0.908).Twelve studies , 34, 35 P = 0.94; Figure P = 0.92). Four studies [P = 0.66; sign test: P = 0.04). Egger's tests showed that there was no bias for these meta-analyses .Analyses of the association between aneuploidy and age revealed no statistical significance. From eight , 26, 33  studies , 18, 34 P < 0.00001; Figure P = 0.001) and OS  compared with diploid ones, based on a random-effects model.Patients with aneuploid tumors had significantly worse DFS , and in ER− BCs than ER+ ones  based on a fixed-effect model. The pooled HR for OS was 0.87  based on the fixed-effect model. The pooled HR for DFS was 0.76  based on the random-effects model. There is no indication of biases for these subgroup analyses .Subgroup analysis was carried out and focused on node-negative BC. Pooled results showed that aneuploidy remained significantly more prevalent in ≤ 2 cm tumors compared to > 2 cm ones . In tumor stage analysis, after excluding Wyss-Desserich et al.'s study [et al. (2013)'s data [P = 0.06). In tumor grade analyses, after omitting Gazic et al.'s study [vs. G1: RR = 2.12, 95% CI, 1.88 to 2.39; P < 0.00001; fixed-effect model; G3 vs. G2: RR = 1.37, 95% CI, 1.28 to 1.46; P < 0.00001; fixed-effect model). Sensitivity analyses of the other outcomes yielded similar results. Egger's test and an exhaustive literature search conferred a substantial degree of confidence in our pooled findings.Sensitivity analyses were performed for all the outcomes. Particularly, the sensitivity test excluding the largest study  showed t's study , which w's study , which uDNA content analysis has been suggested to assess cell kinetics insofar as the DNA percent of S phase cells is identified by histogram. However, the S phase assessment is usually hampered by the overlap between aneuploid tumor and diploid non-malignant cell populations, few cells, background debris, and multiploid/hypertetraploid tumors. For this reason, in human tumor cells, DNA content analysis is mainly used to evaluate the occurrence of aneuploid cell population, an abnormality known to represent the tumor cells. Though the role of aneuploidy as an independent prognostic parameter in BC has been widely studied during the last decades, the results remained controversial. This study is the first meta-analysis focusing on this particular aspect of aneuploidy in BC. It shows that the frequency of DNA aneuploidy is significantly associated with certain prognostic factors, such as tumor size, grade, LN metastasis, and ER status in BC patients. Aneuploidy is more frequently detected in BCs with larger diameters, poorly differentiated cells, LN metastasis, and negative ER expression. However, it was not significantly correlated with menopausal status or age. This study also supports aneuploidy as a significant prognosticator for DFS and OS in BC. Patients with diploid tumors benefited from a significantly reduced risk  for cancer recurrence and death.et al. [et al. [et al. [It is well described that cancer develops from normal tissue through adenoma to carcinoma, and finally metastasis. Aneuploidy has been suggested to be related to cell proliferation and poor differentiation but not disease stage . Our anaet al.  demonstret al. , 39. The [et al.  investig [et al.  also car− BC, especially node-negative ER− BC, which implies that, adjuvant chemotherapy should be selected for this subgroup of patients. And for those with ER+ tumors, appropriate hormone therapy could be applied.There is a close correlation between DNA ploidy and ER status. Due to the biological and clinical significances of DNA ploidy, it should be focused on during therapeutic planning. Systemic chemotherapy is recommended for the majority of BCs . Howeveret al. [The outcomes of BC patients might be affected by treatment modalities. However, this analysis demonstrates that patients with aneuploid tumors perform worse than those with diploidy ones, even though they are supplied with adjuvant therapy. In 2005, Moureau-Zabotto et al.  carried et al. –46. The Besides, DNA ploidy analysis using the DNA flow cytometry method has its advantages. Most molecular/genomic technologies are not applied in routine practice , 48. TheThe strengths of our analyses lie in the thorough literature research and qualified statistical approaches, which provided the veritable results. Besides, all the selected studies are of relatively high quality. However, this meta-analysis has some limitations. First, it is only based on published but not individual patient data. Second, the studies included were not randomized studies. Third, not all outcome measures were reported by all enrolled studies. Hitherto, the results should be taken into consideration with cautiousness. Further studies are required in this field.In conclusion, this meta-analysis reveals that DNA ploidy status might refine the outcome assessment and personalized treatment choice. And a significantly higher frequency of aneuploidy at advanced tumor stages implies an increased genomic instability during BC progression. This finding may have important therapeutic implications in BC. Furthermore, it should be noted that the value of aneuploidy for BC prognosis needs to be validated in further multicenter studies with larger samples and longer follow-up periods. This would be, undoubtedly, a giant step forward to improve the personalized medicine in BC management.th 2015, using the following keywords: ‘DNA ploidy/aneuploidy’, ‘cytometry’, and ‘breast/mammary cancer/carcinoma’. The American Society of Clinical Oncology annual meeting abstract was also retrieved. The language limitation was not applied during search.This meta-analysis was guided by the Preferred Reported Items for Systematic Reviews and Meta-Analysis (PRISMA) statement issued in 2009 . The elee.g., chemotherapy, radiotherapy, hormone therapy); 4) clear definition of DNA diploidy and aneuploidy; 5) the Kaplan-Meier (K-M) method was used to estimate survival rate according to the DNA content, and a Cox proportional hazard regression model was occupied to investigate the relative strength and independent prognostic value of the variables. Alternatively, the study reported hazard ratio (HR) with a 95% confidence interval (CI).To be considered eligible for this meta-analysis, the relevant clinical studies were carefully selected based on the following criteria: 1) available baseline statuses of enrolled women; 2) estimation of overall survival (OS), disease free survival (DFS), or relapse free survival (RFS)-DNA content relationship as aim of the study; 3) patients were diagnosed BC by histological method and the specimens were collected before any anticancer treatment  status, DFS, and OS. DFS events included local and distant recurrences of the original cancer, second primary BC, and death. Hazard ratios (HRs) with 95% confidence intervals (CIs) were applied where possible. If HRs and 95% CIs were not directly reported, we extracted data from K-M curves by Engauge Digitizer 4.1 and then calculated the indexes.I2 index, which describes the percentage of total cross-study variation due to heterogeneity rather than chance, to obtain a quantitative measure of the degree of inconsistency in the results of studies. The fixed-effect model was adopted if no heterogeneity was present (2Pχ < 0.100 and 2I < 50%). When there existed excessive heterogeneity, data were first rechecked for validation. If the heterogeneity persisted, the DerSimonian random-effects model was employed [P values were two-sided.Statistical analyses were carried out following the recommendations of the Cochrane Collaboration Guidelines. The quality of included studies was assessed by the Newcastle-Ottawa Scale (NOS) scale. Fixed-effect or random-effects model chosen according to data heterogeneity was adopted to calculate the pooled estimate  using the Mantel-Haenszel's or the inverse variance method based on the available data type. Sign tests were performed to assess the significance of evidence in each dataset. Cochran's Q statistic was used to check the homogeneity assumption. We calculated the Higgins' employed , 54. Theemployed . Sensiti


P=0.199). There was a threshold approximately five cups of coffee per day, and the inverse association for colorectal cancer appeared to be stronger at a higher range of intake. However, a nonlinear association of rectal cancer with coffee consumption was not observed (P for nonlinearity = 0.214). In conclusion, coffee consumption is significantly associated with a decreased risk of colorectal cancer at ≥ 5 cups per day of coffee consumption. The findings support the recommendations of including coffee as a healthy beverage for the prevention of colorectal cancer.A meta-analysis was performed to assess the association of coffee consumption with colorectal cancer and to investigate the shape of the association. Relevant prospective cohort studies were identified by a comprehensive search of the PubMed, Embase and Web of Science databases from their inception through August 2015. Either a random-effects model or fixed-effects model was used to compute the pooled risk estimates when appropriate. Linear and nonlinear dose-response meta-analyses were also performed. Nineteen prospective cohort studies involving 2,046,575 participants and 22,629 patients with colorectal cancer were included. The risk of colon cancer was decreased by 7% for every 4 cups per day of coffee (RR=0.93, 95%CI, 0.88-0.99;  Colorectal cancer is the third most commonly diagnosed cancer in males and the second in females, with an estimated 1.4 million cases and 693,900 deaths occurring in 2012 worldwide . It has Certain foods and drinks, including coffee, have been implicated in the etiology of colorectal carcinoma , 10. CofIn a previous meta-analysis conducted by Li et al. , the assThe search identified 1211 articles from the PubMed, Embase, and Web of Science databases. After the initial screening, based on the titles and abstracts, 317 articles remained for further full-text assessment. After retrieving the full-text review for detailed evaluation, 19 prospective cohort studies –31 exami2I=41.40%; P=0.031).Nineteen cohort studies –31 inves2I=34.30%; P=0.082; Figure P for nonlinearity=0.004), without a statistically significant association below 5 cups/day. Compared with people who had no daily consumption of coffee, the RR of colorectal cancer directly estimated from the cubic spline model was 1.00  for 1 cup/day, 1.00  for 2 cups/day, 1.00  for 3 cups/day, 0.98  for 4 cups/day, 0.96  for 5 cups/ day, 0.93  for 6 cups/day, 0.90  for 7 cups/day, and 0.87 for 8 cups/day.Seventeen studies , 30, 31 2I=29.90%; P=0.124).Sixteen studies , 30, 31 2I=23.00%; P=0.199; Figure P for nonlinearity <0.001). Compared to people who had no coffee consumption, the RR of colon cancer estimated directly from the cubic spline model was 0.99  for 1 cup/day, 0.98  for 2 cups/day, 0.97  for 3 cups/day, 0.94  for 4 cups/day, 0.91  for 5 cups/day, 0.89  for 6 cups/day, 0.86  for 7 cups/day, and 0.82  for 8 cups/day.Fifteen studies , 30, 31 2I=13.00%; P=0.308; Fifteen studies , 30, 31 2I=11.00%; P=0.333; P for nonlinearity=0.222).Fourteen studies , 30, 31 P > 0.05 for each). Importantly, we found that among studies published before 2000, coffee consumption was significantly associated with the risk of colorectal cancer , but in later studies, coffee consumption was not significantly associated with colorectal cancer risk . According to stratified analyses by the subtypes of coffee, a significant inverse association between decaffeinated coffee consumption and colorectal cancer risk was identified , and caffeinated coffee consumption was not significantly associated with colorectal cancer risk .The results from subgroup analyses that examined the robustness of the primary results and explored the sources of potential heterogeneity are shown in Findings with fixed-effects inverse-variance weighting were similar for colorectal  and rectal cancer  in the analysis of high versus low intake of coffee. In contrast, with fixed-effects inverse-variance weighting, coffee consumption was inversely associated with colon cancer .2I=32.40%, P=0.091) to 1.00 . Similarly, in the dose-response analysis, the non-significant inverse association of coffee intake with colorectal cancer was still unchanged by removing one study at a time, with a pooled RR ranging from 0.95  to 0.99 .In the sensitivity analysis of high versus low coffee intake, omitting one study in turn, no particular study explained the results found for colorectal cancer; the pooled RR ranged from 0.96 , similar to the original analysis including all studies.In addition, the effect of excluding studies from the dose-response analysis on the results was explored. When the analysis of high versus low coffee intake was restricted to the studies that were included in the dose-response analysis, the summary RR was 0.97  after the exclusion of this study. In the dose-response analysis, when this study was excluded, the nonlinear associations of colorectal and colon cancers with coffee consumption did not change  when this study was excluded. Similar results were found for an increment of 4 cups/day coffee consumption and the risk of colorectal, colon and rectal cancers after the removal of this study. Thus, it is relatively appropriate to include this study in our dose-response analysis.The lowest reference category according to one study by Hartman et al.  was lessP=0.944, Egger's P=0.764), colon cancer , and rectal cancer . Additionally, no significant evidence of substantial publication bias was observed for the dose-response analyses .In the analyses of the highest compared with the lowest category, the Begg's and the Egger's tests did not provide any significant evidence of publication bias for studies that investigated the relationship between coffee consumption and colorectal cancer  have been described as having genotoxic and mutagenic properties. For example, coffee contains acrylamide and caffeine, which have potential carcinogenic effects , 35. In We observed a significant inverse association between coffee consumption and colorectal cancer risk in earlier publications (2000 or earlier). One potential interpretation of this finding was that coffee brewing methods had changed over time, and currently, the filter method is more popular, effectively replacing unfiltered forms of coffee, such as boiled coffee, which were more widely consumed by participants in earlier studies . Boiled A significant inverse association with colorectal cancer risk was observed with decaffeinated coffee consumption, but not with caffeinated coffee consumption, which is an interesting phenomenon. There were several potential explanations. The differences could reflect chance or perhaps residual confounding, if the lifestyles differed between individuals who regularly drank decaffeinated coffee and those who drank caffeinated coffee. Previous studies showed that participants who drank decaffeinated coffee tended to drink less alcohol, eat more fruit and vegetables, and consume less red meat than participants who drank caffeinated coffee , 28. AnoThis meta-analysis has several strengths. Because we based the analysis on prospective studies, the findings are unlikely to be explained by recall and selection biases. Our meta-analysis included a large number of studies and more than 22,600 cases, with over two million participants in the coffee consumption analysis. Thus, we had adequate statistical power to analyze the association of coffee consumption with colorectal cancer risk, clarify the shape of the dose-response relationship between coffee consumption and the risk of colorectal cancer, and to detect low to moderate reductions in risk. We quantified the association between coffee consumption and the risk of colorectal cancer by carrying out linear and nonlinear dose-response analyses and found a significant nonlinear association of coffee consumption with colorectal and colon cancer risk.Several limitations should be acknowledged. Most original studies used FFQ to assess levels of coffee consumption. Although validation studies showed that FFQ was a reasonable tool to assess the intakes of coffee, measurement error was inevitable. Second, although the original studies included in our analysis were adjusted for multiple major risk factors for colorectal cancer except for three studies, the possibility of residual confounding by imprecisely or unmeasured confounders should be considered, as coffee consumers tended to follow less healthy behaviors, including lower levels of physical activity, higher prevalence of smoking and overweight/obesity, and higher intake of alcohol and red and processed meat , 22, 32.In summary, this meta-analysis suggests a nonlinear association between coffee intake and colorectal cancer. There was a threshold approximately 5 cups of coffee consumed per day and further reductions in the risk of colorectal cancer with higher intake. Studies with larger sample sizes and longer follow-up times are warranted to confirm our results. Given that coffee is consumed very commonly and the levels of morbidity and mortality of colorectal cancer are high worldwide, the results of our study provide practical and valuable insight into the prevention of colorectal cancer and the study of its etiology.This systematic review was conducted following the Meta-analysis of Observational Studies in Epidemiology (MOOSE) guidelines . We perfStudies were included in this meta-analysis according to the following criteria: (1) the study was a prospective cohort study, including nested case-control studies with a prospective design; (2) the exposure of interest was coffee consumption, including total coffee, caffeinated coffee, or decaffeinated coffee; (3) the outcome of interest was the risk of colorectal cancer; (4) the participants were free of colorectal cancer at baseline; and (5) the study reported risk estimates with corresponding 95% CI for the association between coffee and colorectal cancer or provided corresponding data to calculate the variance. In cases where multiple publications from the same study were available, we included the publication that presented the results with sufficient detail to be incorporated into dose-response analyses or the publication with the largest number of cases.Two investigators (Y. G. and L. Q. L.) independently extracted the data from each study using a standardized electronic format, including the last name of the first author, publication year, study location, sample size, sex, age range or mean age at entry, length of follow-up, number of cases, method of assessment of exposure, outcome measurements, RRs with corresponding 95% CIs for all categories of coffee consumption, and covariates adjusted for in the multivariable analysis. We extracted risk estimates with the most adjustment (when available). If a study only reported caffeinated coffee consumption instead of total coffee consumption, caffeinated coffee consumption was also included in the total coffee consumption analysis. For dose-response analysis, when studies reported the consumption in milliliters per day or week or month, we standardized all the data into cups per day; 237 ml of coffee was equal to 1 cup. Differences in data extraction between the two investigators were resolved by discussion with the third investigator (Z. X. L.).Quality assessment was performed according to the NOS . The scaRRs were considered as the common measure of the association between coffee consumption and colorectal cancer risk, and the hazard ratios (HRs) were considered to be equivalent to RRs. We preferentially pooled multivariable adjusted RR estimates where such estimates were available. When the adjusted estimates were unavailable (three studies), we pooled the unadjusted estimates. A random effects model was used to calculate the summary RRs and 95% CIs for the highest versus lowest level of consumption of coffee and for the dose-response analysis. We compared these findings to the fixed-effects inverse-variance weights in sensitivity analyses. For studies that reported results separately for colon and rectal cancer or for men and women, we combined the estimates using a fixed-effects model to obtain an overall estimate for colorectal cancer or both sexes combined in the primary meta-analysis .For the dose-response analysis, we used the previously described method  to calcuP value for nonlinearity was calculated by testing that the coefficient of the second spline transformation was equal to zero against the null hypothesis [We presented the dose-response results in the forest plots for an increase of 4 cups per day. We also examined the potential non-linear associations between coffee consumption and colorectal cancer risk using the two-stage random-effects dose-response meta-analysis. This was conducted by modeling consumption with the use of restricted cubic splines with 3 knots at fixed percentiles  of the distribution . A P valpothesis .2I statistic (ranging from 0% to 100%). 2I values of 25%, 50%, and 75% represented cut-off points for low, moderate, and high heterogeneity, respectively [Statistical heterogeneity across studies was assessed using the ectively .Subgroup and meta-regression analyses by sex, study location, cancer subsite, duration of follow-up, specific dietary assessment method, subtypes of coffee, publication year, and adjustment for confounding factors such as BMI, smoking, alcohol, physical activity, dairy products/calcium intake, energy intake, fruit and vegetable intake, folate intake and red and the consumption of processed meat were conducted to investigate the potential sources of heterogeneity among studies. Sensitivity analyses were performed to evaluate the effect of removing a single study from the analysis on the pooled risk estimates.P < 0.10. All statistical analyses were conducted with STATA version 12.0 . P values were two-tailed with a significance level of 0.05, except where otherwise specified.The Begg's test  and the


Multiple studies have reported evidence of an inverse association between parity and bladder cancer risk. However, a comprehensive and quantitative assessment of this association has never been conducted. We conducted this study to clarify this issue.Systematic search of PubMed and Embase was performed to identify all the studies. Studies were selected based on strict screening with inclusion and exclusion criteria. Summary relative risks (RR) with 95% confidence intervals (CI) were calculated by using a fixed-effect model, and the generalized least squares trend estimation was employed to compute study-specific RR and 95% CI per live birth increase. Heterogeneity and publication bias were also evaluated.Twelve studies  were eligible in this meta-analysis. The pooled RR of bladder cancer for parous versus nulliparous women was 0.76 (95% CI: 0.70–0.82). Results were similar in the studies that adjusted for BMI, cigarette smoking , and age . The dose-response meta-analysis showed a lower bladder cancer risk  for each live birth increase in parous women. No evidence of publication bias or significant heterogeneity was detected in the above-mentioned analyses.The finding from current meta-analysis suggest that parity may be related to decreased risk of bladder cancer.The online version of this article (doi:10.1186/s12885-016-3023-5) contains supplementary material, which is available to authorized users. Bladder cancer (BC) is one of the most common genitourinary malignancies in the United States, with an estimated 76, 960 newly diagnosed cases and 16 390 deaths in 2016 [Although the exact biological mechanisms for the gender disparity in BC development and progression are not well established, some studies have suggested that sex hormones and their receptors may contribute to this difference –5. DurinTherefore, we find it necessary to further assess the relationship between parity and BC risk by conducting a meta-analysis of current epidemiologic studies and providing a quantitative dose-response analysis.This study protocol was approved by the institutional review board at the West China School of Medicine before initiation and don’t need ethical standard statement.We searched online databases MEDLINE and EMBASE for the studies assessing the relationship of parity and BC risk before Sep 2016, using the following terms with every possible combination considered: parity, pregnancy, reproductivity, fertility, bladder, genitourinary tract, cancer, carcinoma, tumor. We searched the references of all eligible publications to identify additional relevant publications. Studies were included if they met the following criteria: (1) the study had a prospective or case-control design; (2) the study evaluated the relationship between parity and BC risk; (3) data were provided or would allow the calculation of relative risk (RR) or odds ratio (OR) and a 95% confidence intervals (CI). Exclusion criteria were as follows: (1) was not involved with the associations between parity and BC risk; (2) non-human studies, reviews, and comments; (3) studies not reporting primary outcomes; (4) studies based on overlapping patients. When duplicate publications were identified or data in separate publications originated from the same cohort, we used the publication with the largest number of cases and the most applicable data.Data were independently extracted from the included studies by two investigators (YJB and XMW), and disagreement was settled by discussion with the third investigator. The following data were extracted from each study: first author’ name, publication year, country of study, participant demographics, sample size, duration of follow-up or study period, parity number categories, corresponding OR/RR (with their 95% CIs) for each category, and variables adjusted for in the analysis.The methodological quality assessment was conducted for each of the included studies using the Newcastle-Ottawa Scale quality assessment  on three2 statistics [P < 0.10 as evidence of heterogeneity. An I2 score exceeding 50% is considered to indicate the presence of heterogeneity. The random effect model was used to provide summary estimations if there was heterogeneity, or else the fixed effect model was used. Heterogeneity was explored by stratified analysis. Publication bias was explored using Begg’s and Egger’s test [P < 0.05 indicated the existence of publication bias.Given that bladder cancer is a rare disease, the OR was assumed to be the same as the RR and we therefore report all effect sizes as RR for simplicity. We calculated summary RRs and 95% CIs associated with parity and BC risk, if the study considered nulliparous as a reference. Between-study heterogeneity was evaluated with Q and Iatistics . For ther’s test , 17 and P value for nonlinearity was computed by testing the null hypothesis that the coefficient of the second spline was equal to zero. Statistical analyses were performed with Stata version 12.0 software .We carried out a dose-response meta-analysis using the method proposed by Greenland et al.  and OrsiThe article selection process is shown in Fig. 2 = 0%, p = 0.795, Fig. 2 = 0%, p = 0.562). For women who have ever smoked, the summary RR was 0.90  for ever versus nulliparous, without heterogeneity .All included studies in present study investigated the association between ever parity and BC risk. The summary multivariable-adjusted RR of BC for ever versus nulliparous was 0.76 , without heterogeneity (I2 = 17.7% (P = 0.295). The second group (3–4 births) contained seven reports, compared with nulliparous, the overall RR of BC was 0.79 (95% CI 0.68–0.91) with I2 = 0% (P = 0.997). The third group (≥5 births) contained six reports, compared with nulliparous, the overall RR of BC was 0.76 (95% CI 0.66–0.88) with I2 = 0% (P = 0.994).Six cohort and six case-control studies investigated possible association between parity and BC risk. The results of the effects of different parity number on BC risk are presented in Table 2 = 12.79%, P = 0.7832). Compared with nulliparous individuals, the pooled RRs (95% CI) of BC were 0.84 (0.76–0.94), 0.76 (0.67–0.87), 0.74 (0.65–0.85), 0.75 (0.66–0.85), and 0.76 (0.65–0.88) for 1, 2, 3, 4, and 5 live births, respectively (Fig. Dose-response from nine studies , 20–23 sNo evidence of a significant publication bias was observed in our analyses as assessed using the Egger’s test or Begg’s test and no asymmetry was seen in the funnel plots of included studies.To avoid the influence of cigarette smoking, we evaluated the influence of smoking status by adjustment in the eight reports that considered smoking. However, when data were stratified by adjustment for smoking, we did not find a significant difference between summary RRs that were adjusted and those that were not adjusted for smoking Table . MoreoveNumerous epidemiological studies have focused on BC risk factors in order to explain the gender disparity in BC incidence, while there is also a great interest in identifying factors that affect the risk of BC among women. Although previous studies have reported on the associations between parity and BC, to the best of our knowledge, no other earlier studies have clearly documented a dose-response pattern between parity number and BC risk. In present study, we conducted a dose-response meta-analysis to characterize the association between parity and BC risk. Findings from the present study indicated that parous was significantly inversely associated with BC risk compared with nulliparous, especially for non-smokers. Overall, the risk of BC decreased by 24% for women who had given birth and 53% for those who had never smoked and were ever parous.Although the exact mechanisms underlying this inverse association between parity and BC risk are not completely established, several potential mechanisms have been proposed. First, during pregnancy, estrogen and progesterone levels increase drastically and induce substantial alterations to the structure, function, and histology of the bladder , possiblSecond, smoking habits may have influenced the incidence of BC in women with parity. It is well known that prenatal cigarette smoking exposure may cause detrimental effects on reproductive health . TherefoThird, structural changes in the pelvic organs and pelvic floor organization following pregnancy may induce a decreased risk for BC in parous women. Childbirth is an established risk factor for lower urinary tract dysfunction, such as increased urinary frequency, stress incontinence, and overactive bladder , but theMeta-analyses of observational studies are prone to confounding, selection, and information bias as the included observational studies and can present particular challenges that may distort the results. However, our study has some strengths. First, we searched 2 electronic databases, which ensured a broad scope, and we included seven cohorts and seven case-control studies that involved a total of 6,214 cases and 2,693,350 non-cases. This large sample size should have provided sufficient statistical power to detect this potential relationship. Second, to control misclassification, we conducted a dose-response meta-analysis. This is the most comprehensive meta-analysis evaluating the associations between parity and BC. Third, no publication biases were detected in present study and outcome may be unlikely to remain unreported as studies with complex assessment of exposure.This current dose-response meta-analysis indicates that women with parity have an inversely associated BC risk compared with nulliparous women. The exact mechanism underlying this protective effect requires further investigation.


Vascular calcification is a risk factor for the pathogenesis of cardiovascular disease and mortality in dialysis patients. Nevertheless, the association between cardiac valve calcification (CVC) and the outcome of dialysis is still illusive. The purpose of this meta-analysis is to evaluate the association between theCVC and cardiovascular or all-cause mortality in dialysis patients.Literatures involving the baseline CVC and cardiovascular or all-cause mortality in dialysis patients were searchedfrom the PubMed, Embase, as well as two Chinese databases (i.e. Wanfang and CNKI databases). Articles published before November 2016were eligible to the study.Ten studies involving 2686 participants were included. CVC was correlated with increased risk of cardiovascular mortality  and all-cause mortality . Subgroup analysis showed an excess risk of all-cause mortality  among patients with one CVC, and increased risk of all-cause mortality in patients with two CVCs .CVC is correlated with higher cardiovascular and all-cause mortality risk in dialysis patients. Regular follow-up monitoring of CVC may be helpful for risk stratification of patients underwent dialysis. Cardiovascular calcification, aprevalent condition at all stages of chronic kidney disease (CKD), is one of the major causes for increased cardiovascularmorbidity and mortality world wide . It has Vascular calcification has been acknowledged as a strong predictor for cardiovascular morbidity and mortality inCKD population . A crossMeta-analysis was performed according to the recommendations proposed by the Preferred Reporting Items for Systematic Reviews and Meta-analyses Statement . We searStudies met the following criteria were eligible: 1) original observational cohort studies; 2) involving participants with end-stage kidney disease underwent dialysis; 3) evaluating the association between the presence and extent of CVC at baseline and subsequent cardiovascular or all-cause mortality; and 4) reporting outcomesas raw data or risk estimation of cardiovascular or all-cause mortality. In cases of several publications from the same study group, only the most recent complete publication was selected. Studies of cross-sectional design, published in types of review or protocol, or duplicated publication were excluded from the study.Two authors (Z Wang and HY Chen) independently collected data from the eligible studies, and did the data extraction including first author’s name, publication year, research design, geographical region of study, demographic characteristics of patients, examination methods, prevalence of CVC, number of death events, adjusted hazard ratio (HR) and 95% confidence intervals (CI), length of follow-up, and adjustment for covariates. A profound discussion was held in cases of any discrepancies during the data collection until consensus. The Newcastle–Ottawa Scale (NOS) for cohort studies was applied to assess the quality of methodology of each study . A study2 statistic were utilized to evaluate the heterogeneity among studies. Random effect model as described by DerSimonian and Laird was applied in cases of significant heterogeneity (P < 0.10 and I2 > 50%). Otherwise, a fixed-effect model as described by Mentel-Haenszel was applied  that was significantly lower than the other studies. This may be related to the exclusion of heart failure subjects, which may be an important cause for the mortality. For the sensitivity analysis, the heterogeneity showed significant decrease after excluding the study reported by Varma et al. (I2 = 12.4%), which may be related the higher heterogeneity as the study was of low quality. Meta-regression analysis revealed region, follow up duration, dialysis modality, being a multicenter study or not, a randomized study or not, a blinded follow up or not were not the major sources for the inter-study heterogeneity . Besides, the number of physiciansanalyzing echocardiographic recordings served as variable, which also showed no statistical differences (P = 0.054). As only a few studies were included, it may be a source for the heterogeneity despite a P value of > 0.05. For further analysis, subgroup analysis was performed using the number of physicians as the variable, which revealed significant decrease in the heterogeneity. Thus, single physician may increase the bias, which overestimated the effects of CVC on the all-cause mortality.The inter-study heterogeneity showed decrease after grouping based on the hemodialysis (I2 = 0.0%). Only one study  was performed in the non-Asian region, which may be responsible for the heterogeneity. This may be related to the exclusion of patients with heart failure that was an important cause for cardiovascular mortality. Meta-regression analysis revealed region, follow up duration, dialysis modality, being a multicenter study or not, a randomized study or not, a blinded follow up or not were not the major sources for the inter-study heterogeneity . Besides, the number of physiciansanalyzing echocardiographic recordings served as variable (P = 0.061). As only a few studies were included, it may be a source for the heterogeneity despite a P value of > 0.05. For further analysis, subgroup analysis was performed using the number of researchers as the variable, which revealed significant decrease in the heterogeneity . Thus, single physician may increase the bias, which overestimated the effects ofCVC on the cardiovascular mortality. In a previous study, Panuccio et al. (2004) excluded the patients with heart failure, which may increase the heterogeneity. Meta-regression analysis showed heart failure may be a source for the heterogeneity (P = 0.064). In the meta-analysis excluding such study, the heterogeneity showed significant decrease . As heart failure was one of the major causes for the cardiovascular mortality, exclusion of heart failure may underestimate the effects of CVC.For the analysis of cardiovascular mortality, the regions were divided into Asia and non-Asian regions. The heterogeneity in the 5 studies performed in Asia was low (ICVC isan active pathobiological process resulting in calcium deposition involving aging, inflammation, CKD, diabetes mellitus, and phenotypic switch of vascular smooth muscle cells. In an environment with changed profiles of calcification-regulating humoral factors including calcium and phosphate, VSMCs can transform to osteo/chondrogenic phenotype. In addition, other factors such as degradation of elastin fibers also played important roles insubsequent calcification .CKD progression may trigger various abnormalities in mineral and bone metabolism such as hyperphosphatemia, hypercalcemia, secondary hyperparathyroidism, bone disorders, and cardiovascular calcification. To date, several factors were identified as the risk factors of cardiovascular calcification, including metabolic disorder of calcium, phosphorus, parathyroid hormone, and fibroblast growth factor 23, inflammation, oxidative stress, declining of calcification inhibitor such as fetuin-A and matrix Gla protein, as well as pharmacological interventions such as active vitamin D and calcium-based phosphate binders. These factors may increase the cardiovascular events and mortality finally.CVC refers to a condition with the presence of bright echoes of more than 1 mm on one or more cusps of the aortic valve, mitral annulus or mitral valve based on the guidelines of the American Society of Echocardiography . GeneralOur study had some limitations. Firstly, some eligible studies were of small sample size and short follow-up duration, which may lead to limited generalizability. Secondly, the adjust covariates in different studies were not the same, which might affect the results of this meta-analysis. Thirdly, there are some negative findings that are not published, which may affect the results of this meta-analysis. Finally, the result based on the limited studies may be not robust, especially in the subgroup analyses.In summary, this meta-analysis indicates that CVC is associated with higher all-cause and cardiovascular mortality in dialysis patients. The number of calcified cardiac valve is correlated with the mortality risk. Our study provides the evidence that detection of valve calcification is beneficial for risk stratification of dialysis. In future, more prospective trials with large sample size are needed to further evaluate the relationship between CVC and the outcome of dialysis patients.


P < 10−5). The pooled analyses also revealed that higher serum TSH level were significantly associated with the size of TC nodule and malignancy as well as lymph node metastasis. Furthermore, significantly increased THS levels were observed preferentially for papillary thyroid cancer when stratified by histological type of tumors. However, the diagnostic value of TSH level for TC might be limited. These results suggest that higher serum TSH concentration is associated with an increased risk of thyroid cancer.Thyroid stimulating hormone (TSH) is the major growth factor for thyrocytes, but the pathogenic role of serum TSH in thyroid cancer (TC) is unknown. The association between TSH level and the development of thyroid cancer has been widely evaluated recently. However, the results remain conflicting. To develop an understanding of the relationship between TSH exposure and thyroid cancer, a meta-analysis of 56 studies involving 20227 thyroid cancer cases and 50003 controls with benign thyroid nodule was performed. Overall, significantly increased TSH level was observed in thyroid cancer patients compared with controls (RoM: 1.44, 95% CI: 1.32–1.56,  Thyroid cancer (TC) is the most common endocrine malignancy , which iThyroid stimulating hormone (TSH) is a glycoprotein, which stimulates secretion of thyroid hormones, maintenance of thyroid-specific gene expression, and gland growth , 9. SuppOur primary search generated 8,918 citations. After duplicates were removed 5484 citations remained for title and abstract screening. Results from the literature search and study selection process were shown in Figure P < 10−5). In different ethnicities, the pooled RoM for East Asian TC patients was 1.39 , 1.72 for Caucasian cases  and 1.49 for the other ethnic populations  and 1.33  in studies with small and moderate sample size, respectively. When analysis was restricted to the 12 studies with at least 500 TC cases, which should be less prone to selective publication than smaller studies, yielded an pooled RoM of 1.33 . Paediatric thyroid cancer appears to be unique and quite different from the adult form. Significant increased serum TSH levels were both found in adults cases  and paediatric TC patients . In the stratified analysis by study design, a significant increased TSH was detected among studies using retrospective cross-section design and studies using other design  compared with patients with smaller one (Table P < 10−5). There was significant difference in TSH levels between TC cases with and without lymph node metastasis . Given the biological differences between the histological types of TC, subgroup analyses by histological types found that serum TSH levels were significantly increased among PTC (P < 10−5). However, no significant differences in TSH levels were detected in FTC (P = 0.95).The data on TSH level of TC patients stratified by tumour size, TNM stages and lymph node metastasis status were available in 8, 7, and 10 studies respectively Table . When tuP < 10−5); while the OR for TC with TSH between 1.35 and 2.3 was 1.84 , was used as reference group and a TSH of 4.2 or greater was considered above the upper end of normal TSH . The risTo investigate the diagnostic value of TSH for thyroid cancer, a diagnostic meta-analysis was conducted. Many studies showed the distributions of cases and controls according to the TSH range. To explore the optimal cut-off value, true positives, false positives, false negatives, and true negatives for each individual study were calculated according to different cut-off value. The pooled specificity, sensitivity and area under the SROC were summarized in Figure P < 0.05). As the formal test for heterogeneity may not be powerful enough, meta-regression was used to evaluate potential effect modifiers by including ethnicity, sample size, study design, mean age, sex distribution and study quality as covariates. In meta-regression analysis, it was found that sample size  and ethnic population  was potential source of heterogeneity; while study design (P = 0.90), mean age of cases (P = 0.18) and controls (P = 0.79), sex distribution among cases (P = 0.09) and controls (P = 0.76), and study quality (P = 0.05), did not significantly explain between-study heterogeneity. Furthermore, Galbraith plot analyses of all included studies were used to assess the potential sources of heterogeneity. Five studies were found to be the contributors of heterogeneity . One recent study [n = 14) yielded large variance in the effect estimate; removing this study led to almost the same summary RoM . However, omitting the study by Wang et al. (14), which included large sample size and heterogeneous results, did substantially influence the summary RoM . Sensitivity analysis indicated that the results of this meta-analysis were stable, with RoMs and 95% CIs ranging from 1.39  to 1.45   .P = 0.39 by Egger regression test; P = 0.06 Begg adjusted rank correlation test, We found no evidence of small study effects , positive association was maintained regardless of the size of tumor. This suggests that the pathophysiology of TSH in tumorigenesis was similar among TC regardless of tumor size. Therefore, relatively high serum TSH concentrations may not be useful as a malignancy predictor in the assessment of thyroid nodules. When patients with TC were grouped according to TNM, we found that mean serum TSH levels were significantly higher in those with advanced stage disease when compared with those with more localised (stage I and II) disease. A similar finding was present in patients with lymph node metastasis with respect to those with no evidence of node metastasis. These results of advanced disease further suggest that TSH might be involved in the pathogenesis or progression of TC. Stratification of tumors by histological subtype indicated that TSH increased, preferentially for PTC. The reason for the observed tumour-specific difference in the TSH level is unknown. However, different carcinogenic processes may be involved in the genesis of various well-differentiated thyroid cancers because of the presence of different concentration of serum TSH. As studies for FTC are currently limited, further studies with increasing number of FTC cases are warranted to confirm our findings.In the diagnostic meta-analysis, we explored the diagnostic performance based on different TSH cut-off values. The areas under SROC were quite similar when different cut-off values were used, while the specificity and sensitivity depended on the TSH cut-off value. However, the overall diagnostic accuracy was poor. No cut-off value can pass the clinical useful test, which indicated TSH alone is not sufficient for confirmation or exclusion of TC. TSH together with other factors may increase the diagnostic accuracy.TSH stimulates the production and release of thyroid hormones and promotes thyroid cancer growth as well as, invasion, and angiogenesis . TrophicLimitations also inevitably existed in this meta-analysis. Firstly, our results were based on unadjusted; while a more precise analysis should be conducted after adjustment for important confounders . Secondly, data from the studies exhibit statistically significant heterogeneity. Lack of individual-level data prevents us from making efficient assessment of the sources of heterogeneity. Thirdly, the subgroup analysis of associations between TSH level and tumour size, TNM stage, and lymph node metastasis as well as histological type were performed on the basis of a fraction of all the possible data to be pooled, so selection bias may have occurred and our results may be overinflated. Larger studies with detailed histopathological information are needed to confirm our findings. Fourthly, independence of TSH for TC risk was not investigated in current meta-analysis due to the limit access to individual data. However, many studies included in our-meta-analysis have proved its independence by logistic regression analysis , 43. FinIn conclusion, our meta-analysis of 56 studies indicate that a significant increased serum TSH level were associated with risk of TC. Direct evidence from future prospective study is warranted to clarify a cause-and-effect relationship between TSH and TC as well as its diagnostic value.We performed this analysis in accordance with the guidelines of the PRISMA  statement . EpidemiThe outcome of interest was histologically or pathologically confirmed thyroid cancer. Control subjects were defined as thyroid nodule patients or benign surgical patients without thyroid cancer. Eligible studies had to meet all of the following criteria: (1) investigated preoperative serum TSH levels and the risk of TC using either prospective (nested case-control or cohort study) or retrospective design ; (2) original human studies with independent data; (3) no medications that would specifically suppress thyroid function and consequentially TSH response ; (4) serum TSH were measured separately for cases and controls using a reliable assay; (5) the results were expressed as, or could be estimated into, mean and standard deviation (SD). Major exclusion criteria were: (1) overlapping data, (2) case-only studies and (3) insufficient data.For association studies with inconsistent results, the methodological quality should be assessed by appropriate criteria to limit the risk of introducing bias into meta-analyses. A procedure known as ‘Newcastle–Ottawa Scale (NOS)’ has been used to assess the quality of association studies . NOS scoTwo investigators independently reviewed each eligible article and extracted relevant information. The following data were collected from each study: first author, publication date, diagnostic criterion, study design, age, gender, ethnicity, quantified method of serum TSH, sample size, TNM (tumor node metastasis) stages , tumor size , lymph node metastasis , histological subtypes, the mean and SD values of serum TSH concentration among cases and controls, the distributions of cases and controls according to different TSH range.. When only the median and range were reported, we used a conversion formula  to conveI2 statistics (I2 = (Q − df)/Q) to reflect between-study heterogeneity. The percentage of I2 < 25, 25–50, and > 50 indicates low, moderate, and high heterogeneity, respectively [P = 0.05 level.As serum TSH concentrations were measured using different methods across different studies, a ratio method was used to express the difference in mean TSH between TC patients and controls for each study. In brief, ratio of the mean (RoM) was defined as the mean value of the TC case group divided by that of the control group , 83. Thuectively . Sourcesectively , 88. Theectively . Small sectively  and Eggeectively . All the