AhR-related gene may exert pathological influences on sexual differentiation of the brain, causing autistic traits. To ascertain the relationship between AhR-related gene polymorphisms and autism susceptibility, we identified genotypes of them in patients and controls and determined whether there are different gene and genotype distributions between both groups. In addition, to clarify the relationships between the polymorphisms and the severity of autism, we compared the two genotypes of AhR-related genes  with the severity of autistic symptoms. Although no statistically significant difference was found between autism spectrum disorder (ASD) patients and control individuals for the genotypic distribution of any of the polymorphisms studied herein, a significant difference in the total score of severity was observed in rs2228099 polymorphism, suggesting that the polymorphism modifies the severity of ASD symptoms but not ASD susceptibility. Moreover, we found that a significant difference in the social communication score of severity was observed. These results suggest that the rs2228099 polymorphism is possibly associated with the severity of social communication impairment among the diverse ASD symptoms.Exposure to environmental chemicals, such as dioxin, is known to have adverse effects on the homeostasis of gonadal steroids, thereby potentially altering the sexual differentiation of the brain to express autistic traits. Dioxin-like chemicals act on the aryl hydrocarbon receptor (AhR), polymorphisms, and mutations of  Reports of the incidence of autism spectrum disorder (ASD) – characterized by two core symptoms: communication and social deficits and fixed or repetitive behavior  – have bp-dioxin (TCDD), polychlorinated dibenzo-p-dioxins and dibenzofurans (PCDD/Fs), and some polychlorinated biphenyls (PCBs), act on the aryl hydrocarbon receptor (AhR)  theory is one of the leading hypotheses for explaining the mechanism of ASD , 11. Theor (AhR) , 15. Recor (AhR) , 17. Theor (AhR) .AhR)-related gene polymorphisms.Epidemiological studies have shown that PCB exposure at low levels can exert adverse clinical and subclinical effects on sociocognitive functions –23. DepeAhR-related gene polymorphisms and reproductive system diseases, such as endometriosis or infertility  Val189Val as potential candidates  contribute to ASD susceptibility and/or severity.Numerous studies have investigated the association between ertility –27, becaertility . Variousndidates –27. Althorphisms . TherefoAhR-related gene polymorphisms and ASD susceptibility, we identified these genotypes in patients and controls and determined whether there are different gene and genotype distributions between the two groups. Second, to clarify the relationships between the polymorphisms and the severity of ASD, we compared the genotypes of AhR-related genes with the severity of ASD symptoms using the Childhood Autism Rating Scale (CARS)  . Finallye (CARS) . We alsoal cases , 33.Ninety-five children and adults with ASD participated in the present study. Participants with ASD were recruited in two different geographical regions, Tokyo and Nagasaki, in Japan. The participants in Tokyo consisted of 68 ASD patients  and were recruited from the outpatient clinics of the Department of Psychiatry in the University of Tokyo Hospital. The participants in Nagasaki consisted of 27 ASD patients , and they were recruited from two day-care facilities in Nagasaki prefecture for patients with developmental disorders. In both areas, the diagnoses were made by two or more senior pediatric-psychiatric clinicians through structured interviews and reviews of clinical records according to the DSM-IV criteria . Among tAdditionally, 527 adults  were recruited as control participants from the nearby community around the University of Tokyo Hospital, without any psychiatric disorder disturbing their work function. The Mini-International Neuropsychiatric Interview (MINI)  and otheThe race/ethnicity of all participants was Japanese. Potential participants were also excluded if they had any history of substance abuse, recent substance use, head trauma with loss of consciousness, significant fetal exposure to alcohol or drugs, perinatal or neonatal complications, and neurological disorders or medical conditions.The present study was approved by the Ethical Committees of the University of Tokyo Graduate School of Medicine and the Nagasaki University Graduate School of Biomedical Sciences. All participants or parents of the affected individuals provided written informed consent prior to their participation in this study. The experimental protocol was conducted in accordance with the Declaration of Helsinki.The severity of ASD was assessed on the basis of the Japanese version of the CARS . The assAhR codon 554 in exon 10  and ARNT codon 189 in exon 7   in set B. All participants were genotyped by real-time polymerase chain reaction (PCR) analysis using Roche LightCycler 480 II  for the following two single nucleotide polymorphisms (SNPs): 2066853) –28. ReacAhR-related gene polymorphism and susceptibility to ASD. Next, analysis of variance (ANOVA) was used to compare the severity of ASD among AhR-related gene polymorphisms. Third, to assess the severity corresponding to several core behavioral symptoms of ASD, factor analysis with Varimax rotation for CARS was performed, and the factor score was calculated by regression method for each subject. Finally, ANOVA was also used to assess the effects of AhR-related gene polymorphisms for each severity of discriminative behavioral symptoms of ASD identified by the prior factor analysis. The chi-squared test, factor analysis, and multinomial logistic regression analysis were performed using IBM SPSS 20.0 for Windows . The ANOVA was performed using Anovakun software . Comparisons between genotype groups did not demonstrate statistically significant differences with regard to children’s age, sex, or IQ level. To investigate the relationship between each AhR-related gene polymorphism and susceptibility to ASD, the chi-squared test was used to evaluate the genotype distribution according to developmental status. No statistically significant association was observed between any of the polymorphisms and susceptibility to ASD.The genotype and allele frequencies of AhR-related gene polymorphism and the severity of ASD, one-way ANOVA was used for the total CARS score of ASD patients as the dependent variable and the genotype of each AhR-related gene polymorphism as the independent variable. There was a statistically significant difference between genotype groups for ARNT polymorphism (rs2228099), as determined by ANOVA . Holm’s sequentially rejective Bonferroni post hoc test revealed that the total CARS score of the GG genotype was statistically significantly higher than those of the GC genotypes , whereas there were no statistically significant differences in score between the CC genotype and both genotype groups . Additionally, there were no statistically significant differences between genotype groups for AhR polymorphism (rs2066853) [F = 0.26]. Taken together, these results suggest that ARNT polymorphism modified the severity of ASD among the AhR-related genes examined in the current study.Next, to investigate the relationship between each AhR-related genes and the severity of each of the three behavioral symptoms of ASD.To assess the severity according to several core behavioral symptoms of ASD, factor analysis with Varimax rotation was performed for the CARS data. The analysis produced three factors with eigenvalues greater than one. These factors accounted for 55.1% of the common variance. Table AhR-related gene polymorphism and the severity of several of the core symptoms of ASD, one-way ANOVA was performed for the factor score of CARS of ASD patients as the dependent variable, with the genotype of each AhR-related gene polymorphism as the independent variable. Similar to the result for the total score, there were no statistically significant associations between factor scores and genotype groups for AhR polymorphism (rs2066853) (Table ARNT polymorphism (rs2228099), a significant difference was observed for the factor score of the “social communication” factor, but not for the “sensory and emotional response” factor or “stereotypies” factor. Holm’s sequentially rejective Bonferroni post hoc test revealed that the factor score of the GG genotype was statistically significantly higher than that of the GC genotypes , whereas there were no statistically significant differences in score between CC genotype and both genotype groups  . Although the exact molecular and physiological mechanisms underlying this effect remain unknown, a recent study suggested that silent mutations may contribute to mental disorders  may have provided a better index (ARNT gene polymorphism to aspects of social communication.Several limitations of the present study should be noted and taken into consideration in future studies. First, the main limitation is the relatively small patient group. We did not observe a significant association between genetic SNPs and susceptibility to ASD in this study, and one possible explanation is the small sample number, as a large number of subjects are needed for case–control studies based on the frequency distribution. Therefore, studies involving a larger number of subjects are essential to generalize our results. Second, the number and age of subjects between the cases and control groups were not well matched. It is possible that the risk modulation by er index . This heer index . TherefoARNT GG genotype had more severely impaired social communication than those with GC genotype in ASD, indicating that the differences in social functioning in ASD patients may be modulated by ARNT variants. Considering that ARNT is a component of AhR cascades, vulnerability to environment chemicals, especially dioxin-like chemicals may affect the severity of impaired social communication, although the functional analysis of ARNT gene polymorphism remains to be performed. To identify the neuronal mechanism underlying this effect, combining the present experimental paradigm with neurophysiological indicators of brain activities is warranted in future studies.In conclusion, the current results showed that individuals with the TF and SN were involved in conducting the experiment, analyzing and interpreting data, and drafting the article. RI, JM, and CK were involved in recruiting the participants and diagnosing the participants with ASD. MT, TS, and NK were involved in conducting the experiment, analyzing and interpreting data, and revising the article. KS conceived of the study, participated in its design and coordination, and drafted the manuscript. All the authors have read and approved the final manuscript.The authors declare that the research was conducted in the absence of any commercial or financial relationships that could be construed as a potential conflict of interest.


Observational studies suggest greater height is associated with increased ovarian cancer risk, but cannot exclude bias and/or confounding as explanations for this. Mendelian randomisation (MR) can provide evidence which may be less prone to bias.We pooled data from 39 Ovarian Cancer Association Consortium studies . We applied two-stage predictor-substitution MR, using a weighted genetic risk score combining 609 single-nucleotide polymorphisms. Study-specific odds ratios (OR) and 95% confidence intervals (CI) for the association between genetically predicted height and risk were pooled using random-effects meta-analysis.Greater genetically predicted height was associated with increased ovarian cancer risk overall (pooled-OR (pOR) = 1.06; 95% CI: 1.01–1.11 per 5 cm increase in height), and separately for invasive  and borderline  tumours.Women with a genetic propensity to being taller have increased risk of ovarian cancer. This suggests genes influencing height are involved in pathways promoting ovarian carcinogenesis. The degree to which these factors could account for the observed association is uncertain. Mendelian randomisation (MR) uses genetic markers as proxies for environmental exposures and, due to the singular qualities of genotype data, can provide complementary evidence by overcoming many biases affecting conventional studies.5 We used MR to examine the relationship between height and ovarian cancer risk in the Ovarian Cancer Association Consortium (OCAC),6 aiming to provide more certainty about the relationship between height and ovarian cancer risk. We hypothesised that greater genetically predicted height would be associated with increased risk.Observational studies have reported a positive association between adult height and ovarian cancer risk.7 Participants were genotyped via the Collaborative Oncological Gene-Environment Study.8 Twenty-two studies provided height data  use, education and age at menarche information) for >50% of their participants. We first considered all cases, then stratified by tumour behaviour. Secondary analyses stratified by histologic subtype/behaviour.We pooled data from 16,395 genetically European women with primary ovarian/fallopian tube/peritoneal cancer and 23,003 controls from 39 OCAC studies , 609 remained (92 genotyped/517 imputed) (Supplementary Table The Genetic Investigation of ANthropometric Traits (GIANT) Consortium had previously identified 697 single-nucleotide polymorphisms (SNPs) significantly associated with height.β-coefficients for their association with height reported by GIANT. The score represents predicted additional height conferred by these variants, compared to having no height-increasing alleles. We report results for 5 cm increments.We constructed a weighted genetic risk score (GRS) for height by summing height-increasing alleles across the 609 SNPs (‘GRS-609’/‘the GRS’), weighting alleles by 7 Briefly, we used individual-level OCAC data for two-stage predictor-substitution MR 12: first, we predicted height from the weighted GRS for all participants using coefficients from linear regression in 17,649 controls with height data; second, within each study, we used logistic regression to model disease status on GRS-predicted height. Models adjusted for age and five principal components for population structure.8 We combined study-specific estimates using meta-analysis,13 generating pooled odds ratios (pOR) and 95% confidence intervals (CI) for the trend in risk per 5 cm increase in predicted height. We had 97% power to detect an OR of 1.10 (mRnd tool).14Statistical methods have been described previously.r2 ≥ 0.9 (‘GRS-363’), SNPs representing distinct loci9 (‘GRS-377’), and directly-genotyped SNPs (‘GRS-92’). We examined whether potential confounders of the association in observational studies were associated with the GRS. To assess robustness to pleiotropy (where SNPs may influence risk via pathways not mediated through height), we conducted MR-Egger regression15 and assessed smaller GRSs excluding SNPs with the highest probability of acting via other pathways from GRS to outcome , hormone levels and in/near tumour initiation/growth genes). We identified these potentially pleiotropic, pathway-specific SNPs via the NHGRI GWAS Catalog,16 the UCSC Genome/Table Browsers18 and from lists of SNPs nominated for iCOGS genotyping by ovarian, breast and prostate cancer researchers .Sensitivity analyses included removing 16 SNPs with MAFs <5%, and restricting to SNPs with imputation Secondary analyses defined cases by histologic subtype/behaviour. Among 16 studies with height/confounder data, we conducted conventional analysis  and compared results with MR-estimates among the same women.Analyses were performed using SAS 9.2  and Stata 13.0 . This work and each contributing study was approved by the appropriate institutional review board/ethics committee. All participants provided informed consent.p < 0.0001).We included 16,395 cases  and 23,003 controls , p < 0.001). A 1-unit GRS-609 increase was associated with 5.2 cm greater height. Average height was 6.2 cm greater in the highest vs. lowest GRS quartile.The GRS-609 was normally distributed in controls, ranging from 15.45 to 18.99 . It explained 13% of variance in height, 17% after adjusting for age and principal components  or the simplified GRS-363  and height among controls , consistent with known genetic overlap between these traits.20Cochran’s p = 0.1; MR-Egger beta = 0.163 corresponded to an OR per 5 cm of 1.13 (95% CI: 1.02–1.25), confirming a significant positive association).Women with greater genetically predicted height had a modestly increased risk of developing ovarian cancer  Fig. ; Table 2In contrast, for women with height and confounder data (16 studies), the conventional analysis suggested no association . Conducting MR within the same 16 studies yielded results similar to overall analyses   and low-grade/borderline serous cancers   and/or early-life exposures .Observational studies are subject to biases  which cannot be ruled out as possible explanations for observed associations. By using genotype, the MR technique can overcome some of these biases, given three assumptions. We confirmed the two verifiable assumptions: the GRS was associated with height, and not with most known confounders. The GRS-menarche age association is unlikely to explain the observed association, because age at menarche is only weakly associated with ovarian cancer, and women with later menarche have if anything lower ovarian cancer risk, so if this affected our results, we would expect the true effect to be at least as strong as the reported association. Also, removing hormone-related SNPs, or adjusting for menarche age, did not attenuate estimates. Owing to the limited current biological understanding of all 609 SNPs, we could not conclusively exclude the presence of alternate pathways from height genes to ovarian cancer (assumption three). However, MR-Egger and sensitivity analyses excluding pathway-specific SNPs provided some evidence for their absence, minimising the likelihood that our observed association is explained by pathways separate from height/growth. Although height data were not available for the entire population, this is unlikely to have affected our results as we used these data only to refine the height predictions from the GRS, and there is no reason to believe the GRS-height relationship would be different for women with and without height data. Further strengths of our analysis include the large number of SNPs and power to detect modest differences.4 The lack of association seen in the OCAC conventional height analysis reflects the greater potential for bias in case–control studies and illustrates the value of MR in overcoming these biases. Few previous studies have examined borderline cancers separately, a strength of our analysis. Previous observational studies have not reported consistent patterns by histologic subtype23; our secondary analyses were under-powered to resolve this question.Despite potential limitations of conventional observational studies, our MR-estimate is almost identical to previously reported associations, suggesting previous estimates were not appreciably biased. The World Cancer Research Fund/American Institute for Cancer Research meta-analysis of 24 prospective studies, and a study pooling 47 prospective/case–control studies, both reported a significant 7–8% increase in risk (combining invasive/borderline cancers) per 5 cm height increase.Using MR, we have established that the previously observed association between height and ovarian cancer risk is unlikely to have been explained by bias, and that genetic factors influencing height play roles in ovarian cancer development. Height could therefore be used, with other risk factors, to identify women at elevated risk. Further research should continue to explore mechanisms underpinning this association.Supplementary Information


Transmembrane 6 superfamily member 2 (TM6SF2) E167K variant, which causes hepatocellular fat retention by altering lipoprotein secretion, on liver damage and metabolic traits in chronic hepatitis C patients is still debated. We performed a systematic review and meta-analysis to clarify this relationship. Four studies with a total of 4325 patients were included. The risk of histologically-determined advanced steatosis, fibrosis, and cirrhosis (but not of severe inflammation) were increased in carriers of the TM6SF2 variant (P < 0.05). Unlike the inconsistent association with steatosis severity, due to the confounding effect of infection by the genotype-3 hepatitis C virus, the TM6SF2 variant was robustly associated with advanced fibrosis  and in particular with cirrhosis . Regarding metabolic features, individuals positive for the TM6SF2 variant exhibited 5.8–12.0% lower levels of circulating triglycerides and non-HDL cholesterol (P < 0.05). Carriers of the variant were leaner, but there was high heterogeneity across studies (I2 = 97.2%). No significant association was observed between the TM6SF2 variant and insulin resistance or hepatitis C viral load (both P > 0.05). In conclusion, the TM6SF2 E167K variant promotes the development of steatosis, fibrosis and cirrhosis in patients with chronic hepatitis C. Conversely, this variant reduces circulating atherogenic lipid fractions.The impact of  However, in patients with chronic hepatitis C (CHC) there is a wide inter-individual variability in the susceptibility to develop of progressive liver disease, which may lead to cirrhosis and hepatocellular carcinoma2.Despite the advent of highly effective direct antiviral agents4. Both viral factors, e.g., infection by HCV-genotype 3 (G3), and host factors, including overweight, type 2 diabetes and at risk alcohol intake, contribute to the development of steatosis in patients with CHC7.Steatosis, defined as hepatocellular fat accumulation exceeding 5% of liver weight, is a typical histological feature of CHC, which is associated with higher likelihood of progression of liver damage and fibrosisPatatin-like phospholipase domain-containing 3 (PNPLA3) gene, the major common genetic determinant of hepatic fat content, influences the development of steatosis, and the progression towards advanced fibrosis and hepatocellular carcinoma in patients with CHC11.Inherited host factors modify the susceptibility to the development of steatosis and liver damage progression in patients with CHC. We and others have previously shown that the I148M variant of the Transmembrane 6 superfamily member 2 gene (TM6SF2), which encodes for the E167K aminoacidic substitution, has been identified as a determinant of hepatic fat content and lower serum lipoproteins levels13. This mechanism is related to the inhibition of lipidation of very-low density lipoproteins (VLDLs)14, which cause lipid compartmentalization within hepatocytes16, thereby triggering steatohepatitis and fibrogenesis18.Recently, the rs58542926 C > T genetic variant of the 21 and fibrosis development21 in this specific population. However, in the CHC setting the association with fibrosis development remains controversial22, and the overall evidence does not support an association of the E167K variant with liver enzymes, and index of hepatic inflammation23.Subsequent studies conducted in patients with CHC suggested that the E167K variant is also associated with steatosisTM6SF2 E167K variant with histological liver damage  in patients with CHC by performing a systematic review and meta-analysis of the available literature. As secondary outcomes, we also evaluated the impact of the E167K variant on circulating lipids, other metabolic traits and their interaction with TM6SF2-histological damage associations.Therefore, this study investigated the association of the The initial literature search yielded 19 potential references after excluding 14 duplicates across three databases. The manual search did not identify any additional studies. Three studies were omitted based on their titles and abstracts, and 12 studies were excluded because of specifically listed reasons. Finally, four eligible studies were selected with high inter-author consistency (Cohen’s kappa = 0.69). The flow chart is presented in Fig. TM6SF2 E167K variant did not violate the Hardy-Weinberg equilibrium (P > 0.05), with carriage of the E167K variant (EK + KK genotypes) frequency ranging from 6.3 to 12.2%. With respect to the liver biopsy results, four studies evaluated the extent of histological steatosis/fibrosis, and three evaluated the extent of inflammation . Compared with HCV-G3 patients and healthy controls, non-HCV-G3 patients had a significantly lower prevalence of E167K variant carriers , with low inter-subgroup heterogeneity . Only Milano M. et al., detected a positive association between the TM6SF2 E167K variant and severe inflammation. However, a sensitivity analysis did not reveal any specific study as a potential confounder that influenced the overall results  without obvious discrepancy , despite the classification using diverse scoring systems  and HOMA-IR .The impact of the y Figure . ConvincTM6SF2 variant was associated with lower BMI . However, after excluding the study by Coppola et al., which had a smaller sample size and inferior study quality, carriage of the E167K variant was associated with a 2% decrease in BMI  in the absence of obvious heterogeneity (I2 = 0%).The TM6SF2 genotype .All four studies compared the HCV-RNA load in subgroups stratified by the e Figure . HCV-RNATM6SF2 variant in high consistence (I2 < 25%). As Figure phenotype-disease ranged across 1; Figure TGs and TC were selected as candidate phenotypes for the Mendelian randomization (MR) analysis of the for stable association with Begg’s funnel plot was applied to examine publication bias Figure . AlthougTM6SF2 E167K variant on liver disease severity in patients with CHC, while also addressing its influence on metabolic and viral features. The major finding are that the TM6SF2 E167K variant favors the development of steatosis and fibrosis, but not of inflammation, in individuals with CHC infection. In our analyses, E167K was associated more robustly with fibrosis than with steatosis. This result might be related to known higher inter-observer variability in the classification of steatosis than fibrosis that is consistently reported across studies28, as well as to the well-established steatogenic effect of infection via HCV-G329. Alternatively, the measurement of hepatic intracellular lipid content via histology might underestimate the disturbance in lipid handling among patients with advanced liver fibrosis because of the reduction in hepatic lipid accumulation with progressive accumulation of extracellular matrix in the liver30.The current study performed a systematic review and meta-analysis of the literature concerning the effect of the TM6SF2 E167K variant conferred a stable and significant predisposition towards the development of progressive hepatic fibrosis. In particular, our pooled results found that the TM6SF2 variant predicted the risk of cirrhosis. Overall, the results of the present analysis suggest that carriage of the E167K TM6SF2 variant predicts an increased risk of clinically significant and advanced fibrosis in patients with CHC. The same conclusion might be drawn for the presence of steatosis in general because histological steatosis is associated with faster fibrosis progression independent of its aetiology31, and other genetic determinants of hepatic fat content,  are associated with fibrosis development patients with CHC32. Because the impact of E167K TM6SF2 on fibrosis progression is not restricted to patients with CHC33, and carriers of the variant have more advanced fibrosis at diagnosis on average, future studies should determine whether this genetic factor can be used to predict the risk of liver-related complications after viral eradication.Importantly, carriage of the TM6SF2 variant was associated with lower circulating serum lipids, primarily on TGs and non-HDL cholesterol. This finding is consistent with the model in which the TM6SF2 variant favours steatosis development because of decreased VLDL secretion and lipidation35. Using an MR analysis as an integrated approach across genotypes, phenotypes, and histological features, we found that the genetic susceptibility conferred by the TM6SF2 variant on steatosis (but not fibrosis) severity was influenced by lipid phenotypes with regard to their causal effects on lipid accumulation in patients with CHC. In fact, 1 unit (mmol/L) of decrease in TG/TC level might increase the susceptibility of steatosis approximately 2-folds. Importantly, the TM6SF2 variation showed a common but distinguished impact on steatosis both in patients with G3 and those with non-G3 HCV infection, suggesting that the TM6SF2 variant independently influences hepatic fat accumulation even in the presence of the strong viral factors that inhibit VLDL secretion via other mechanisms37. Therefore, the dissociation of the E167K variant between more severe liver damage but more favourable lipid profile observed in the general population and non-alcoholic fatty liver disease (NAFLD) patients is also present in patients with CHC.Furthermore, the TM6SF2 genotype and liver damage was not confounded by metabolic cofactors. Indeed, in keeping with observations of the general population22 and patients with NAFLD34, the E167K variant was not associated with an increased risk of hyperglycaemia or insulin resistance. Intriguingly, the current summary of the available evidence revealed an association between carriage of the TM6SF2 E167K variant and slightly reduced BMI. Although it is tempting to speculate that the association might depend on impairments in chylomicrons and VLDL secretion (thereby supporting adipose tissue growth in carriers of the mutation), we must remember that the E167K variant does not affect BMI in the general population12. Conversely, the E167K variant was associated with an increased risk of cirrhosis, an important cause of malnutrition and low BMI, which might have driven the observed association38. Additional epidemiological investigations evaluating single patient data are required to confirm this association.The association between the 16 is that the TM6SF2 variation was not associated with histological hepatic inflammation in CHC. This is possibly due to interference of the E167K variant in the process of VLDL secretion that impairs HCV viremia and infectivity39. This hypothesis is also supported by the evidence collected in genetic studies of individuals with NAFLD, in whom genetic susceptibility to hepatic lipid accumulation is the major determinant of liver damage progression41. One important caveat to note is that steatosis and TM6SF2 variation increase the risk of lobular necroinflammation with the infiltration of granulocytes34, a typical feature of non-alcoholic steatohepatitis24. Conversely, histological scores developed for CHC primarily capture the portal and peri-portal infiltration of lymphocytes, and can miss the nonalcoholic steatohepatitis (NASH)-related features of liver damage possibly associated with TM6SF2 variation. Nevertheless, our findings reinforce the notion that circulating aminotransferase levels, which are not affected by TM6SF2 variation in patients with CHC23, are not an accurate predictor of the severity of damage and fibrosis.One important difference between the results of the present analysis and findings derived from studies conducted in patients with NAFLD42. Nevertheless, IPD meta-analyses of studies, are still needed to reach even more reliable conclusions less subjected to potential bias. Fourth, a lack of adjustment for steatogenic factors such as alcohol intake and less consistent data clusters in the subgroup analysis might have biased the pooled results. Finally, despite the uneven distribution of TM6SF2 genotypes across patients infected by diverse HCV genotypes  in individuals of European descent with CHC infection. This variant is associated with a reduction in circulating non-HDL cholesterol and TG levels. Decreased TC and TG levels might amplify the genetic susceptibility of TM6SF2 variant on advanced steatosis.In conclusion, by systematically reviewing and meta-analysing the available literature, we found that the TM6SF2 variants on hepatic steatosis and relevant complications in patients with CHC was conducted using the Medline, EMBASE, and the Cochrane Library databases without language restriction . Additionally, we manually searched for “grey” literature on website  according to the guidance of the Cochrane Handbook for Systematic Reviews . The reference lists of enrolled studies and publications with citations of included papers were also reviewed for suitable papers. Our literature search strategy is shown in Table A comprehensive search for literature addressing the genetic associations of TM6SF2 E167K variant was genotyped using reliable methods; 3. Histological features were assessed via liver biopsy under the guidance of a pre-defined scoring system; 4. The risk of hazardous variants on the susceptibility of hepatic histological lesions was reported or could be calculated. If two (or more) studies included the same cohort, then the most recent study was included.Qualified studies were selected when they met the following criteria: 1. Patients were clearly diagnosed with CHC but were free of other severe liver/systemic disease; 2. The TM6SF2 genotyping . Then, additional information concerning the association between the TM6SF2 E167K polymorphism and specific histological features  including the scoring system, number of patients with different disease severities classified by TM6SF2 variants, comparisons, corresponding odds ratios (ORs), and adjusted covariates were also collected. In addition, quantitative indicators of lipids , low density lipoprotein cholesterol [LDL-C), high density lipoprotein cholesterol [HDL-C)), insulin resistance , and viral load of HCV RNA, categorised by TM6SF2 variants, were collected when available. The corresponding authors of the original studies were contacted via e-mail for data missing from the published papers when necessary.The following information provided by the enrolled studies was extracted: study name, population characteristics ), study design, HCV genotype distribution, and 44. Eleven items were included in the qualitative checklist  study guidelinesst Table .vs. EE) to evaluate the genetic impacts, because of low prevalence of carriers with the homozygous variant (KK). Quantitative phenotypes that used different units across studies were unified before comparisons. For covariates reported in participants with CT and TT genotypes, the data were combined according to the number (N), mean and standard deviation (SD) using pre-defined formulas45.We adopted a dominant model (EK + KK http://digitizer.sourceforge.net/). Reported medians and interquartile ranges were transformed to mean values and SDs according to pre-defined methods46. The OR (if not given) and deviation of genotype distribution taken from the Hardy-Weinberg equilibrium were tested using the chi-square test.If the original results were not reported, then data were extracted using Engauge Digitizer version 4.1  method48. We separately calculated the pooled risk stratified by potential confounds to assess their impacts on the associations between the TM6SF2 variant and histological features. A cumulative meta-analysis was performed to identify the change in effect as the study sample sizes increased. A sensitivity analysis was performed by omitting one study sequentially to examine its effect on the overall results49. Publication bias was evaluated using Egger’s funnel plot and Begg’s test50.The difference in quantitative phenotypes was evaluated via the pooled standard mean difference (SMD) and corresponding 95% confidence intervals (CIs). With respect to the categorical variables, pooled ORs and 95% CIs were used to assess the genetic effects on disease predisposition. Pooled ORs and SMDs were calculated using MetanTM6SF2 E167K polymorphism and histological severity (primarily steatosis and fibrosis) were examined via a Mendelian randomisation (MR) approach in absence of reverse causation and unadjusted confounds52. In our study, exposure indicators, which were more strongly correlated with the TM6SF2 variant in low heterogeneity  were selected for the MR analysis. The risk associated with carrying the TM6SF2 variant on more severe steatosis/fibrosis was defined as OREK/KK vs. EE, and the mean difference on the quantitative phenotype was defined asΔP. ORphenotype-disease was calculated using OREK/KK vs. EE1/ΔP as a non-confounded effect with confounders on the risk of more severe steatosis/fibrosis per 1 unit of elevation on the candidate phenotype.The causal effects of the quantitative phenotypes on the genetic associations between the 2 test. Thresholds of 25%, 50%, and 75% were considered as representing low, moderate, and high heterogeneities, respectively53. A fixed-effect model was used in the case of nonsignificant heterogeneity , and a random-effects model was used for other conditions. All analysis was performed by STATA 12.0 . P < 0.05 was considered significant.The heterogeneity caused by discrepancies across different studies was assessed using the ISupplementary files


Circulating miR-31 was found to be associated with cancers detection and prognosis. The present meta-analysis aimed to explore the effect of circulating miR-31 on cancer detection and prognosis.The studies were accessed using multiple databases. RevMan5.3, Meta-DiSc 1.4, and STATA14.0 were used to estimate the pooled effects, heterogeneity among studies, and publication bias.A total of 14 studies with 1397 cancer patients and 1039 controls were included. For the 12 prognostic tests, the adjusted pooled-AUC was 0.79 (95% CI: 0.73-0.86) as the pooled sensitivity, specificity, positive likelihood ratio (PLR), negative likelihood ratio (NLR), diagnostic odd ratio (DOR) from 10 tests was 0.79 (95% CI: 0.76-0.82), 0.79 (95% CI: 0.76-0.82), 3.81 (95% CI: 2.90-5.01), 0.26 (95% CI: 0.20-0.35), and 16.81 (95% CI: 9.67-29.25), respectively. For the 5 prognosis analyses, the pooled HR (hazard ratio) of overall survival (OS) was 1.55 (95% CI 1.30-1.86) for high versus low circulating miR-31 expression. However, high expression of circulating miR-31 did not significantly increase the risk of poor differentiation  and LNM  in lung cancer.Circulating miR-31 is an effective biomarker and could be used as a component of miRs signature for cancer detection and prognosis surveillance. MicroRNAs  are short, single stranded RNA molecules, which primarily bind messenger RNAs (mRNAs) at 3′UTRs via partial complementarity with the “seed sequence”. MiRNAs MiR-31, a highly evolutionarily conserved miRNA, plays an important regulating role in embryonic implantation, development, bone and muscle homeostasis, and immune function . AbundanFourteen eligible studies included in the meta-analysis Figure , 11 fromIn the 14 eligible studies, 10 studies presented the AUC (area under receiver operating characteristic curve) of circulating miR-31 on cancers detection/diagnosis with 1122 cases and 1039 controls, although only 8 studies had the values of sensitivity and specificity for the diagnosis tests. In addition, the participants in 2 studies were divided into two groups for testing and validation. Consequently, we assessed the overview quality of 12 diagnosis tests and reported them in 2I=99%, QP < 0.0001), the random-effects model was used to calculate the pooled effect. As shown in Figure QP = 0.770), the pooled sensitivity, specificity, positive likelihood ratio (PLR), negative likelihood ratio (NLR), diagnostic odd ratio (DOR) was conducted with the value being 0.79 (95% CI: 0.76-0.82), 0.79 (95% CI: 0.76-0.82), 3.81 (95% CI: 2.90-5.01), 0.26 (95% CI: 0.20-0.35), and 16.81 (95% CI: 9.67-29.25), respectively , 0.71 (95%CI: 0.57-0.86), 0.88 (95%CI: 0.82-0.93) for lung cancer, colorectal cancer, and esophageal cancer, respectively. The pooled sensitivity, specificity, PLR, NLR, and DOR of subgroup were presented in Table 2 = 65%, QP = 0.06; for LNM: I2 = 87%, QP < 0.0001), the random-effects model was used. High expression of circulating miR-31 did not significantly increase the risk of poor differentiation  and LNM  in lung cancer (There were 3 and 4 studies reported the cases of poor differentiation and the cases of lymph node metastasis (LNM) by circulating miR-31 level, and all of them were the lung cancer patients. Because of significant heterogeneity among the studies , the fixed-effects model was used. The pooled HR (hazard ratio) of OS was 1.55 (95% CI 1.30-1.86) for high versus low circulating miR-31 expression  by circulating miR-31 level for 612 cancer patients. Because of no heterogeneity . In addition, each of the 9 tests was also excluded sequentially; the summary sensitivity and specificity, PLR, NLR, and area under sROC were altered (data not shown). For the cancer prognosis, each of the 5 studies was sequentially excluded; high miR-31 expression still significantly increased the risk of OS throughout (data not shown). All of them indicated that the present pooled estimated were stable.Z = 0.62, Pr> |z| = 0.54). Given this result, we performed Egger's test where evidence of significance publication bias was found . Consequently, we performed trim and fill analysis to adjust the final effect; the adjusted AUC was 0.79 (95% CI: 0.73-0.86) with P < 0.0001 for heterogeneity. Due to circulating miR-31 acting as a diagnostic biomarker of cancer ). Furthermore, the reference for all effects of prognosis (ORs or HRs) was reformatted as low circulating miR-31 expression, and the multivariate analysis effects were chosen for pooled analysis. The quality of diagnostic test studies was assessed by the Quality Assessment of Diagnostic Accuracy Studies 2 (QUADAS2). Newcastle-Ottawa Scale (NOS) was applied to assess the quality of each eligible study in prognosis analysis and the score ≥6 was considered at high quality.2) and Q tests (chi-square test). If no statistical heterogeneity was found , a fixed-effects model was used to estimate the pooled AUC, sensitivity, specificity, PLR, NLR, DOR, OR and HR. Otherwise, a random-effects model was used. Moreover, Begg's and Egger's tests were used to assess publication bias, and trim and fill analysis was used to adjust the pooled effects by STATA 14.0 [P< 0.05 was considered statistically significant.This meta-analysis involved use of Review Manager 5.3 (Cochrane network), Meta-DiSc 1.4, and STATA 14.0. When AUC, HRs and 95% CIs were not provided directly in some studies, Engauge Digitizer 4.1 and Origin 8 were used to analyze AUC as well as HRs and 95% CIs from the ROC and the Kaplan-Meier curve, respectively. The heterogeneity among studies was tested by Inconsistency (IATA 14.0 . All tesThis meta-analysis is the first to demonstrate that the circulating miR-31 has relatively high effect on cancer detection and prognosis surveillance. The expression of circulating miR-31 might be an effective biomarker for surveillance of cancer.


ALDH2) rs671 polymorphism and essential hypertension (EH). However, the results remain inconsistent. This study aimed to clarify the association between ALDH2 rs671 polymorphism and EH susceptibility.Several studies have been conducted to examine the association between aldehyde dehydrogenase 2 family  and alleles . An additional meta-analysis using 4204 cases and 5435 controls established that rs671 was significantly associated with EH (P<0.00001).Significant differences were observed between EH cases and controls at the level of both genotype (χALDH2 rs671 polymorphism and EH susceptibility. In addition, the results of the breakdown analysis by gender suggest a male-specific association between the ALDH2 rs671 polymorphism and EH.The results of our case-control study and meta-analysis showed that there is a significant association between  Essential hypertension (EH) is a critical cardiovascular risk factor that may lead to stroke, coronary heart disease, diabetes and other diseases . In ChinEH is a complex disease, on which both environmental factors and genetic factors have an important impact . AlcoholALDH2) is located on chromosome 12q24 and is one of the key enzymes involved in ethanol metabolism .This population-based case-control study included 1094 EH cases and 1236 healthy controls that participated in routine health examination at local community health centers between April and July 2013 in Yinzhou District, Ningbo City, Zhejiang Province, China. The cases were recruited if they met the following criteria: (1) systolic blood pressure (SBP) ≥140 mm Hg and/or diastolic blood pressure (DBP) ≥90 mm Hg when taking no antihypertensive medication; (2) previously diagnosed with EH; (3) taking antihypertensive medication; and (4) aged 40–70 years. All subjects were free from secondary hypertension, diabetes mellitus, renal disease, thyroid disease or a history of cancer, and were ethnic Han Chinese. The study protocol was approved by the Medical Ethical Committee of the Affiliated Hospital of Hangzhou Normal University. After written informed consent was obtained from the subjects, a face-to-face interview was conducted to collect information, including demographic (e.g. sex and age) and lifestyle (e.g. smoking and drinking) data, and 5-ml blood samples were collected. Those who smoked at least one cigarette per week were defined as current smokers. Those who drank at least once per week were defined as current drinkers. Thus those former smokers and never smokers were classified as non-smokers, and those former drinkers and never drinkers were defined classified as non- drinkers. Besides, weight and height were also measured using standardized methods during the interview and body mass index (BMI) was calculated by the standard formula . A further sex-stratified association showed that the rs671 polymorphism was significantly associated with EH risk in men [AA/AG vs. GG: OR (95% CI) = 0.76(0.58–0.98)] but not in women.As shown in Besides, the analysis of the distribution of drinking habit showed that there was no significant difference between cases and controls (P = 0.583), even after being adjusted by confounding factors (P = 0.358).We found 70 potentially relevant publications by searching the existing literature databases. After applying the inclusion and exclusion criteria, 10 studies , 22–28 wThe detailed information of each study included in the meta-analysis is presented in After combining all qualified data, the total number of cases and controls were 8963 and 13 047, respectively, from eight eligible case-control studies. Overall, a significantly decreased risk was observed under four genetic models: co-dominant model AA versus GG ;co-dominant model AG versus GG ; dominant model AA/AG versus GG ; allelic contrast model A versus G ] , Fig 2. After stratification by sex, the association between rs671 polymorphism and EH risk remained significant in all genetic models overall and in Japanese male subjects Tables  and 5. WPotential publication bias of this meta-analysis was detected by funnel plot , which rALDH2 gene was significantly associated with EH risk in the case-control study and meta-analysis. The genotype frequencies of ALDH2 rs671 polymorphism were 53.7% (GG), 40.3% (AG) and 6.0% (AA) in the cases, and 49.1% (GG), 43.0% (AG) and 7.9% (AA) in the controls in our case-control study. A significant association between the rs671 polymorphism and EH risk can be proved in all genetic models for Chinese cases and controls. This comprehensive analysis of different genetic models improves the accuracy of prediction and reduces the bias of single model prediction. Subgroup meta-analysis by country indicated a decreased risk of EH risk in the dominant model  for Japanese cases and controls. Besides, the association between rs671 polymorphism and EH risk remained significant in all genetic models overall and in Japanese male subjects. We only found a significant association under the allelic contrast model among Chinese male subjects , and no significant association between rs671 polymorphism and EH risk was found in women. Thus, we concluded that the GG genotype of rs671 appears to be associated with an increased risk of EH only in male subjects, especially in Japanese male subjects.Our results showed that the rs671 polymorphism in the ALDH2 is responsible for the metabolism of toxic aldehydes [ALDH2 may have a lower risk of alcohol-induced high blood pressure than people with the wild-type enzyme, who can consume more alcohol without experiencing acetaldehydemia [ALDH2 could reduce ROS-induced vascular contraction in angiotensin-II (AngII) hypertensive mice. And ALDH2 protected both the microvasculature and microvasculature against reactive aldehydes generated under the condition of sustained oxidative stress [ALDH2 rs671 polymorphism was significantly associated with EH risk. Besides, the results showed that alcohol consumption was not a risk factor for EH, for there was no significant difference of drinking habit between cases and controls (P = 0.583). Thus, we concluded that the rs671 polymorphism of ALDH2 gene was likely to be an independent risk factor.Mitochondrial ldehydes .Individuehydemia . Severale stress , 30, 31.A recent meta-analysis was published and also reported that the GG genotype of the rs671 polymorphism was a risk factor for EH . HoweverALDH2 activity, which detoxifies ROS-producing aldehyde adducts [ALDH2 rs671 polymorphism [A study has shown that the rs671 polymorphism is associated with EH risk among Mongolian women but not men . Our res adducts . Meanwhi adducts . In addimorphism .Our study had some limitations. First, all case-control studies and meta-analyses were conducted in China and Japan, therefore, our findings might be applicable only to Asian populations. Second, due to the lack of uniform background data for studies included in meta-analysis, the data were not further stratified by other factors such as age, smoking, alcohol consumption and other lifestyle factors.In conclusion, this study provides evidence that the rs671 GG genotype may influence the risk of EH independently of alcohol consumption, and mainly affects male subjects. Further investigations with larger sample sizes and detailed gene–environment data should be carried out to confirm these results, to provide an evidence base for public health management of EH.S1 Table(DOCX)Click here for additional data file.S2 Table(DOCX)Click here for additional data file.S3 Table(DOC)Click here for additional data file.S1 Text(DOCX)Click here for additional data file.


In sensitivity analysis, the pooled result remained significant  when we restricted blood glucose categories in the range of nondiabetic subjects. We also detected a J-shaped non-linear dose-response relationship between blood glucose concentration and risk of liver cancer. There is evidence that elevated blood glucose increases risk of liver cancer across the range of prediabetes and diabetes. Considering the rapidly increasing prevalence of prediabetes and diabetes, controlling blood glucose may lower the risk of liver cancer.The question of whether elevated blood glucose is a risk factor for liver cancer has been intensively studied, yet with inconsistent results. To explore the relationship between blood glucose concentration and risk of liver cancer, we conduct a meta-analysis of prospective studies. Literature search was comprehensively performed using database of PubMed, EMBASE and the Cochrane Library through October 2016. Random-effect models were used to combine the effect estimations. Eight articles containing ten studies with a total of 1975 liver cancer cases were included. The pooled RRs demonstrated that elevated fasting blood glucose was associated with increased risk of liver cancer  with mild heterogeneity (I Liver cancer is the six most commonly diagnosed cancer and the leading cause of gastrointestinal cancer-related death worldwide, especially in China , 2. The 1 exposure play dominant roles in most high-rate countries, however, in low-rate countries, obesity, overdrinking, hepatitis B virus (HCV) infection and metabolic syndrome are the major risk factors. Diabetes was reported to associate with a significant 101% increase in risk of liver cancer based on a previous meta-analysis of twenty-five cohort studies [A vast number of factors have been recognized to contribute to the incidence, progression and outcome of liver cancer. Interestingly, these identified risk factors also vary between high-rate and low-rate regions . Chronic studies .in vitro studies and epidemiological studies [Hepatocarcinogenesis caused by hyperglycaemia and hyperinsulinemia through upregulating the production and bioavailability of insulin-like growth factor-1 (IGF-1) has been observed in animal models,  studies , 11. How studies . This prThe previous epidemiological studies referring to the association between blood glucose concentration and liver cancer risk yielded conflicting results –20. PartThe review was registered in PROSPERO-international prospective register of systematic reviews (registration number. CRD42016049496). We designed, analyzed and reported this meta-analysis following the Meta-analysis of Observational Studies in Epidemiology (MOOSE) guidelines . We searStudies were included in this meta-analysis if they met the following criteria: 1) had a prospective design (cohort or nested case-control); 2) reported measurements of blood glucose levels; 3) the outcome of interest was liver cancer incidence; 4) adjusted RRs and corresponding 95% confidence intervals (CIs) were provided.A predefined data extraction form was applied to extract data from relevant articles by two reviews (H.H. and T.Z.) independently. The following information was extracted: the first author's last name, year of publication, study location, study design, age, gender, baseline, number of cases, follow-up time, contrast of the most extreme quantiles, estimated effect size, corresponding 95% CI and adjusted factors. Quality assessment was conducted using Newcastle-Ottawa Quality Assessment Scale (NOS) and studies with an NOS score ≥7 were considered high quality .2 statistics to assess heterogeneity across the studies [We calculated the summary RRs of liver cancer for the highest versus lowest level of blood glucose using the random-effect model described by DerSimonian and Laird, which is in consideration of heterogeneity both between and within studies . If stud studies . Publica studies .P-value for non-linearity was calculated by testing the hypothesis that the coefficient of the second spline was different from zero. We assigned the dose in each category using the midpoint or median level of blood glucose. However, if lower boundary of the lowest category was not reported, it was assumed as 3.9 mmol/L (70.2 mg/dL) for the reason that the lower limit of normal fasting glucose was approximately 3.9 mmol/L [For the dose-response analysis, we used the method proposed by Greenland et al., Longnecker et al.  and Orsi9 mmol/L . If the P-value < 0.05 was considered to be statistically significant.Finally, we first performed sensitivity analysis by omitting any specific study to assess the robustness of the pooled results . Secondly, we repeated the analysis by restricting blood glucose categories in the range of nondiabetic subjects (levels below 126 mg/dl) to detect whether the increased risk of liver cancer was caused by diabetes. Subgroup analysis was conducted based on study region, gender, category number, number of cases, follow-up time and control for potential confounders. Stata Version 12.0 software  was used for all analyses, and We identified 3113 studies through a comprehensive search and 3099 studies were eliminated after review of title or abstract . No evidence of publication bias was detected  for per 0.56 mmol/L (10 mg/dL) increase in fasting blood glucose.For the dose-response analysis, we excluded study conducted by Inoue et al. that only reported two categories because at least three categories were needed for the study-specific trend estimation. The pooled RRs for risk of liver cancer were 1.11  after we restricted blood glucose categories in the range of nondiabetic subjects (levels below 126 mg/dl).Sensitivity analysis in our study indicated similar results. The summary RRs of liver cancer risk for the highest vs. the lowest categories of fasting blood glucose remained significant  when we restricted blood glucose categories in the range of nondiabetic subjects (levels below 126 mg/dl). The To the best of our knowledge, this is so far the first meta-analysis to summarize evidence on the relationship between fasting blood glucose and risk of liver cancer. According to the results of our study, elevated fasting blood glucose was significantly associated with increased risk of liver cancer across the range of prediabetes and diabetes. The risk of liver cancer quantitatively increased by 77% in the category of highest blood glucose compared with the lowest. We further found a J-shaped non-linear relationship between fasting blood glucose and risk of liver cancer.As shown by the results of subgroup analyses, the risk of liver cancer increased with the rises in blood glucose concentration in all subgroups except for subgroup that was stratified by gender. Null association was found in female subjects in this study. The gender disparity in liver cancer incidence has been long well described and it may attribute to a higher prevalence of smoking, alcohol consumption, HBV or HCV infection in males. Our results revealed that susceptibility to higher blood glucose in males may also contribute to the gender disparity of liver cancer occurrence. However, because of a limited studies reporting the result and potential lack of statistical power, this result should be cautiously interpreted. Considering study region, the significant results remained stable no matter in high-risk or low-risk areas. Several risk factors have been established for liver cancer, including HBV or HCV infection , obesityDiabetes (defined as FBG >126 mg/dl) was associated with increased risk of hepatocellular carcinoma in both males and females based on a latest meta-analysis of 25 cohort studies. However, the role of prediabetes (defined as FBG 100–125 mg/dl) on incidence of liver cancer risk was still uncertain. Significant association between prediabetes and liver cancer risk was found in three previous studies and null association was found in five studies. In the sensitivity analysis, when the blood glucose categories above 126 mg/dl were excluded, significant associations for elevated blood glucose and non-linear dose-response relationship still existed, which confirmed the hypothesis that prediabetes status increase liver cancer risk.Several possible mechanisms have been proposed to explain the hypothesis between hyperglycaemia and increased liver cancer incidence. Experimental and epidemiological studies have suggested that hyperglycaemia and hyperinsulinemia may upregulate the production and bioavailability of insulin-like growth factor-1 (IGF-1), which could promote cellular proliferation and inhibits apoptosis by receptor-mediated pathways in live cancer cells , 38, 39.Important strengths should be acknowledged in this meta-analysis. All studies met the inclusion criteria had a prospective design. As liver cancer itself would induce diabetes or hyperglycemia, concerns might raise that the association between elevated blood glucose and liver cancer could be partly caused by reverse causality. Under such assumption, there might be a positive feedback loop in exploring the cause-effect relationship between blood glucose concentration and risk of liver cancer. To exclude possible reverse causality from diabetes or hyperglycemia induced by liver cancer, we repeated the analysis by restricting blood glucose categories in the range of nondiabetic subjects and the result was robust. In addition, we used NOS score to assess the study quality and all studies had a high-quality. As the sample size of primary single study was relatively small, our study increased the statistical power to detect possible association between fasting blood glucose and liver cancer risk and further determined a precise risk estimation. Sensitivity analyses in this meta-analysis proved the stability of our conclusions concerning comparison of the highest category versus the lowest category and non-linear dose-response relationship after omitting one study in turn and restricting blood glucose categories in the range of nondiabetic subjects (levels below 126 mg/dl).Potential limitations should be considered. First, exposure measured only at baseline may not reflect blood glucose levels over a long time follow-up. Especially for subjects with hyperglycemia or diabetes, they tended to control glucose by taking antidiabetic medications or improving life style, which couldn't be neglected as bias would be introduced under these conditions. Second, gender-specific risk for female participants was analyzed in only two studies, although both studies indicated null association. This phenomenon should be further examined and confirmed. Third, in spite of some important confounders seemed unlikely to alter the role of fasting blood glucose on the increment of liver cancer risk based on our subgroup analyses of adjustments, other factors potentially accounted for the observed association cannot be ruled out. For example, information with respect to use of antidiabetic medications was not available in most studies. Four, With regards to subtypes of liver cancer such as hepatocellular carcinoma (the most common type), hepatoblastoma, cholangiocarcinoma and other rare ones, the included studies did not provide corresponding effect estimations, respectively. Further studies should place more emphasis on association between blood glucose concentration and less common subtypes of liver cancer and help to identify more risk factors for these kinds of liver cancer.This meta-analysis provides evidence for the hypothesis that fasting blood glucose is significantly associated with increased risk of liver cancer, and this association is dose-dependent. The combined result of increased liver cancer risk begins to be significant as blood glucose concentration is above approximately 6.50 mmol/L (117 mg/dL). Considering the rapidly increasing prevalence of prediabetes and diabetes, controlling blood glucose may lower the risk of liver cancer.


Conflicting evidence exists for observational studies on whether tobacco smoking is a risk factor for diabetic nephropathy (DN) in patients with type 1 (T1DM) and type 2 diabetes mellitus (T2DM). In this meta-analysis, we aimed to assess the effects of tobacco smoking on the development of DN.st, 2017 for cross-sectional, case-control, and prospective cohort studies. We screened reference lists of retrieved articles. Summary relative risks (SRRs) and 95% confidence intervals (CIs) were calculated using a random-effects model.We searched MEDLINE and EMBASE databases from their inception to March 31P = 0.006) and T2DM , respectively. In patients with T1DM, the SRR (95% CI) was 1.25 (0.86–1.83) for microalbuminuria only, 1.27 (1.10–1.48) for macroalbuminuria only, and 1.06 (0.97–1.15) for end-stage renal disease (ESRD). In patients with T2DM, the SRR (95% CI) associated with ever smoking was 1.46 (0.94–2.26) for microalbuminuria only, 1.72 (1.04–2.84) for macroalbuminuria only, and 1.10 (0.36–3.33) for ESRD.A total of nineteen observational studies  were identified, involving more than 78,000 participants and a total of 17,832 DN cases. Compared with never-smokers, there was an augmented SRR (95% CI) of DN in ever-smokers in patients with T1DM (1.31 [1.06–1.62]; Our meta-analysis suggests evidence for cigarette smoking as an independent risk factor for the development of DN in patients with both T1DM and T2DM. The incidence of diabetes mellitus (DM) is increasing worldwide and not only does it affect quality of life for these patients but the condition also results in a huge economic burden for individuals and countries . DiabetiCigarette smoking is undoubtedly one of the leading causes of premature death, predominantly resulting in cardiovascular disease and cancer . With rehttp://www.embase.com/) and MEDLINE  from the inception of these databases to March 31st, 2017. The search terms included the following key words: 1) smoking OR cigarette OR tobacco OR nicotine; 2) proteinuria diabetes OR diabetic macroalbuminuria OR diabetic microalbuminuria OR diabetic nephropathy OR diabetic nephropathies; and 3) cohort OR case-control OR cross-sectional. The references of included studies and relevant reviews were also checked for eligible studies.Two investigators (HF and JN) independently screened the original articles, which were written in English. We comprehensively searched for relevant studies in two databases  or urine albumin excretion rates (UAER). According to the UACR or UAER, DM patients were divided into 3 groups: normoalbuminuria (UAER: < 30 mg/d or UACR: < 30 mg/g), microalbuminuria  and macroalbuminuria (UAER: ≥ 300 mg/d or UACR: ≥ 300 mg/g). Patients in the micro- and macroalbuminuria groups were considered to have DN.Studies were included according to the following criteria: 1) the design was based on an observational study evaluating the association between smoking and the development of DN; 2) the risk estimates and their confidence intervals [CIs] were at least adjusted or matched for age and hypertension; 3) if more than 1 study originated from the same population, the latest or most informative study was included. Animal experiments, chemistry, cell line studies, editorials, commentaries, review articles, and case reports were all excluded. Data on other forms of tobacco use, such as cigar and pipe, were not analyzed. We did not consider the grey literature. Two investigators (HF and JN) independently reviewed all potentially relevant articles to determine whether an article met the general inclusion criteria, and disagreement was resolved by discussion between the two investigators.Two researchers (HF and JN) independently extracted the following data from each publication: the first author’s last name, year of publication, study locations, sample size, gender, age, smoking category, the methods of assessment of outcomes, duration of follow-up in cohort studies, relative risk [RR] and 95% CI, and co-variables that were adjusted for in the analysis. From each study, we extracted the risk estimates that reflected the greatest degree of control for potential confounders.Based on a random-effects model, we pooled study-specific RRs and their 95% CIs from all included studies. This random- effects model was developed by DerSimonian and Laird, which robustly considers heterogeneity among studies . We used2 statistics. The I2 statistic is the percentage of total variation across studies due to heterogeneity rather than through chance. Results were defined as heterogeneous for P-values < 0.10 or I2 was > 50% [In assessing heterogeneity among studies, we used the Cochran Q and Ias > 50% . We explP < 0.10 was considered to be representative of a significant statistical publication bias. We also performed a nonparametric “trim and fill” procedure to further assess the possible effect of publication bias in our meta-analysis. All statistical analyses were performed using STATA, version 11.0 . A two-sided P value < 0.05 was considered to be statistically significant.Publication bias was assessed using funnel plots and the further Begg’s adjusted rank correlation and Egger’ regression asymmetry tests , 31. P <n = 8), the United States (n = 3), Europe (n = 7), and multi-centers (n = 1).Based on the study selection criteria, we identified a total of 664 potentially relevant articles (653 articles from MEDLINE database and 11 articles from EMBASE database). Of the 664 individual papers, 48 papers were thoroughly assessed by reading their full-text. After reviewing the cross-reference list, a further 5 articles were identified. By studying these 53 articles, 34 were excluded for various reasons Figure . The con2 [All articles were published between 1995 and 2016 and involved more than 78,000 participants and a total of 17,832 DN cases. In all studies, relative risk estimates were adjusted for age, gender, and blood pressure. All but one study made adjustments for duration of DM . More thP = 0.006, I2 = 66.9%; Figure P = 0.287, I2 = 19.8%; Figure P = 0.004, I2 = 73.9%; Figure Six cohort studies and 1 cross-sectional study , 22–24 rP = 0.009, I2 = 67.7%). Table As shown in Table P < 0.001, I2 = 71.6%; Figure P = 0.003, I2 = 82.6%; Figure Six cohort studies, 1 case-control study and seven cross-sectional studies , 25, 28 P = 0.374, I2 = 6.7%) and cross-sectional/case-control studies . Subgroup analysis was performed by the outcome of DN: microalbuminuria only , macroalbuminuria only , and ESRD , and by locations: Asia , Europe  and the USA . We also conducted a sensitivity analysis by omitting one study at a time and calculating the SRRs for the remainder of studies, and found that there were no changes in the direction of effect when any one study was excluded (As shown in Table excluded .P = 0.133) and Egger’s (P = 0.466) tests did not reveal evidence of publication bias but visual inspection of the funnel plots revealed significant asymmetry. The trim-and-fill method suggested that 2 additional risk estimates were needed to balance the funnel plot and the summary risk estimates lost statistical significance  test did not reveal evidence of publication bias but visual inspection of the funnel plots and Egger’s test (P = 0.087) revealed significant bias. The trim-and-fill method suggested that 4 additional risk estimates were needed to balance the funnel plot and the summary risk estimates were still statistically significant, albeit much weaker , suggesting that cessation of smoking significantly decreased the risk of the development of DN. Furthermore, the elevated risk of DN in former smokers lost statistical significance. A prospective study by Chuahirun et al.  and PhisIn the meta-analysis of the effects of cigarette smoking specifically on type 1 and type 2 DN, we observed similar summary risk estimations (SRR = 1.31 vs. 1.44), suggesting that cigarette smoking may exert an equally detrimental influence on T1DM and T2DM. When we performed meta-analysis according to different stages of DN, we found somewhat stronger, albeit not significant, risk estimation for microalbuminuria and macroalbuminuria compared with ESRD. These results indicated that the influence of smoking begins in the early stages of diabetes. In addition, cigarette smoking is possibly a weaker risk factor in the development of ESRD compared with hypertension and duration of DM . HoweverTo make our results more reliable, we made concerted efforts in several aspects of the study as follows: (1) this meta-analysis is the first quantitative review of smoking status and the development of DN, which included a comprehensive, systematic search of the literature using a broad search strategy to identify studies. (2) This meta-analysis included 10 prospective studies, which should have eliminated selection and recall bias. (3) Smoking has effects on many factors, including BMI, vascular disease, dyslipidemia, glycemic control and other important physiological function, and most of the included studies were adjusted for potential confounders. (4) We performed subgroup and meta-regression analysis to explore the source of heterogeneity. We suggest that study design might be one of the sources of heterogeneity.However, our study has some limitations, which should be considered. First, many studies were based on a cross-sectional or case-control design, which does not allow for causal inference and can overestimate measures of association. When restricting to prospective cohort studies, a significant, albeit weaker association was found between cigarette smoking and the development of DN in patients with T1DM and T2DM, respectively. Second, as with all observational studies, residual confounding cannot be completely ruled out. For example, smokers tend to be associated with heavy alcohol consumption behaviors . DunklerP = 0.046), accounting for 21.7% of the significant heterogeneity observed among studies.A third limitation of our study was the high heterogeneity across studies that were present, which may be related to the study design, number of cases, method of exposure measurement, definition of DN, and adjustment for confounders. The definition, classification of smokers, and the methods used for quantifying tobacco exposure are different, which may yield varied results with regard to the strength of the association. In addition, the diagnostic criteria for DN varied, with some studies categorizing UACR 30–300 mg/g as microalbuminuria, UACR > 300 mg/g as overt nephropathy, while other studies defined DN as UAER > 30 mg/24 h on any annual evaluation. Considerable heterogeneity could make the results less robust. The meta-analysis of cohort studies on T2DM showed little variability, whereas significant heterogeneity was observed among cross-sectional/case-control studies. The meta-regression analysis suggested that study design was the most significant factor  provided evidence for such bias for studies on T2DM. Further examination using the trim-and-fill method suggested 4 additional risk estimates were needed to balance the funnel plot, and the summary risk estimates were still statistically significant, albeit these were weaker (SRR =1.24).Finally, publication bias is of a concern since the grey literature tends not to be published in MEDBASE or EMBASE databases. This bias may have led to an overestimation of the true association. Indeed, Egger’s test (In conclusion, results from our meta-analysis of observational studies clearly demonstrate an adverse impact of smoking on the development of DN in patients with T1DM and T2DM.


I2 = 53.8%; p = 0.007). High levels of total cholesterol (TC), low-density lipoprotein cholesterol (LDL-C), and triglycerides (TG) were significantly associated with a decreased risk of AMD . The stratified analysis based on AMD subtypes showed that these blood lipids were only significantly associated with the risk of early AMD . The association between the blood lipids and AMD risk did not differ substantially based on the other characteristics of the participants. A high HDL-C level was associated with an increased AMD risk, whereas participants with high TC, LDL-C, and TG concentrations may show a decreased risk for this disease. Further well-designed large studies are warranted to confirm the conclusions.Lipid metabolism may be involved in the pathogenic mechanism of age-related macular degeneration (AMD). However, conflicting results have been reported in the associations of AMD with blood lipids. We performed a meta-analysis including a total of 19 studies to evaluate associations between blood lipids and this disease. The result reported that the high level of high-density lipoprotein cholesterol (HDL-C) obtained with an increment of 1 mmol/L could result in a significantly increase in the AMD risk of approximately 18% (relative risk (RR), 1.18; 95% confidence interval (CI), 1.01 to 1.35;  Age-related macular degeneration (AMD) is a leading cause of irreversible vision loss among people aged 65 and older in western countries . In factAlthough the pathogenic mechanism remains elusive, the deposition of lipoproteins has been implicated in the pathogenesis of AMD and atherosclerosis . A high Therefore, we conducted a meta-analysis pooling the results of all available association studies between these lipids concentrations and the risk of AMD. Furthermore, a stratified analysis was also performed to examine the associations of these lipids with differentiation in AMD subtype.This meta-analysis was performed in accordance with Preferred Reporting Items for Systematic Reviews and Meta-Analyses (PRISMA) guidelines.We conducted a comprehensive search of the PubMed, EMBASE, and ISI Web of Science databases from their inception to January 2016 for relevant studies that estimated the association between serum lipids and risk of AMD using the following search terms: (Cholesterol or lipids or high-density lipoprotein or HDL or low-density lipoprotein or LDL or triglycerides or triglycerides (TG)) and . The search was not restricted to any language. In addition, the reference lists of the retrieved articles were subsequently evaluated to identify any studies not identified from the preliminary literature searches. When necessary, the corresponding authors were contacted for additional information.All of the studies evaluating the associations of total cholesterol (TC), high-density lipoprotein cholesterol (HDL-C), low-density lipoprotein cholesterol (LDL-C), and TG levels with AMD risk were potentially included. The following criteria were included for the selection: cohort, case-control, or cross-sectional studies published as an original article; the major objective of the study included in the meta-analysis was to evaluate the relationship between these blood lipids and any types of AMD; the information on the type of AMD (early AMD or late AMD (wet or dry)) was assessed in the study; and the relative risks (RRs) [or odds ratios (ORs) or hazard ratios (HRs)] with 95% confidence intervals (CIs) or sufficient data to calculate these parameters were reported. Main exclusion criteria for a study were as follows: the study was a review or editorial comment; the study could not provide sufficient information; the study was not regarding AMD research. When multiple publications reported the same or overlapping data, the study with the largest sample size was included in the analysis. Two investigators (Yafeng Wang and Mingxu Wang) independently reviewed the retrieved records. Any inconsistencies were resolved through consensus with a third author (Le Ma) for adjudication.The following characteristics of each of the identified studies were collected: first author, publication year, study name, country, study type, sample size, mean age, sex, sample source, exposure, classification criteria outcomes, diagnosis method, type of AMD, and covariates in a fully adjusted model. Data on different AMD subtypes were extracted; moreover, the RRs of not only total AMD but also the AMD subtypes (early or late AMD) were all abstracted. Furthermore, geographic atrophy (GA) or CNV were also extracted for late AMD. When the studies only provided subcategories of the AMD disease status, the gratings were collapsed into a single AMD group.The quality of the included studies was independently appraised with the Meta-analysis Of Observational Studies in Epidemiology (MOOSE) guidelines . When leI2 test (I2 > 50% was considered high). In cases of potential heterogeneity, a meta-regression was performed to explore the potential sources of heterogeneity between studies. Subgroup analysis was conducted based on study type , ethnicity , classification criteria of AMD [International Classification and Grading system (ICGS) vs. Wisconsin Age-related Maculopathy Grading System (WARMGS) vs. Age-Related Eye Disease Study (AREDS)], type of AMD (early or late AMD (GA/CNV)), mean age of participants (≥65 vs. <65 years), and study quality (high vs. low). In addition, sensitivity analyses were performed, after sequentially removing one study at a time to evaluate the stability of the results. Begg’s test, Egger’s test, and funnel plots were evaluated to assess for publication bias risk [p values less than 0.05 were considered statistically significant, and in Egger’s linear regression and Begg’s rank correlation, a level of 0.10 was used.A RR with 95% CI is a commonly used measure of effect of interest in the medical and public health literature. When HRs and incidence risk ratios were reported, these parameters were directly considered RRs. Because AMD is not common, we therefore reported ORs as equivalent to RRs. The generalized linear models were used in many studies and therefore we assumed a linear relation between exposure and outcome. RRs were expressed for a standardized increase in the lipid concentrations of 1mmol/L which is the most frequently measure. All lipid concentrations were converted, if necessary, to 1mmol/L. For studies that reported results separately by gender, AMD stage, or race, the risk estimates were pooled using a fixed-effects model before the study was included in the overall analysis. According to the heterogeneity, a random-effects model or fixed-effects model was used in our meta-analysis. The heterogeneity among individual studies was evaluated by calculating the Cochran’s Q statistic and the ias risk ,15. We uA total of 3886 articles were retrieved from the search, and 47 potentially relevant studies were eligible for further review. After reviewing the full-text articles, 21 articles 19 studies) were included in this meta-analysis  [8,9,10, studies n = 18), gender (n = 14) and smoking (n = 10), whereas a fewer number of adjusted for body mass index (n = 6), hypolipidemic drug use (n = 3), and alcohol consumption (n = 2).The main characteristics of the studies are described in I2 = 53.8%; p = 0.007; I2 = 26.6%; p = 0.17). Moreover, the results of sensitivity analysis suggested that no single study could influence the overall pooled estimates. Begg’s funnel plot and Egger’s regression test did not reveal any evidence of the presence of publication bias in the eligible studies (both p > 0.1).The relationship between the HDL-C level and risk of AMD was evaluated in 15 studies, comprising 53,981 participants. Among these, most studies showed inverse associations between a higher HDL-C level and AMD risk, and significant associations were observed in seven studies. The results of the present meta-analysis revealed that an elevation in the HDL level of 1 mmol/L increment could result in a significant increase in the AMD risk of approximately 18% , but not on late stage . The subgroup analysis revealed that most of the other characteristics shared consistency in the direction of the effect. In addition, the sensitivity analysis showed that the significant relationships of the pooled RRs remained stable. We did not observe any evidence of publication bias .We evaluated the association of the LDL-C level with AMD in 10 studies with a total of 27,668 participants. With the exception of the study conducted by Colak et al. , all of I2 = 58.9%; p = 0.001; I2 = 46.2%; p = 0.05) for early stage, but not for late stage AMD . No associations were observed for subtypes of late stage AMD . The significant heterogeneity decreased after the stratified analysis by AMD subtypes. The results of subgroup analysis showed that the associations between TC level and risk of AMD did not significantly differ according to these characteristics. Furthermore, we performed sensitivity analyses and confirmed the robustness of the results. No significant publication bias was observed .The association between the TC level and AMD was investigated in 18 studies with a total of 54,862 participants. Most of these studies showed a protective tendency, and two studies showed a significantly increased risk of this disease. The pooled results revealed that the participants with an increase in a high TC level of 1 mmol/L were at decreased risk for AMD  but not for late stage AMD . The results of the subgroup analysis based on these characteristics presented consistent effect directions. The sensitivity analysis showed that the pooled RRs remained stable. There was no evidence of publication bias .Subsequently, the association of TGs with AMD was evaluated in nine studies, with a total of 38,467 participants. The results revealed that an increase in the TG level of 1 mmol/L would significantly reduce the risk of AMD , is complicated and involves multiple genes which may also be associations with AMD. Recently, two genome-wide association studies identified that several genetic loci including rs493258 and rs10468017 (LIPC), rs3764261 (CETP), rs1883025 (ABCA1), and rs12678919 (LPL) were heavily implicated in HDL metabolism and might have the associations with AMD ,42. As tAs another risk factor for atherosclerosis, LDL-C, was found to play a protective role in the development of AMD in the present meta-analysis. The possible mechanism of action for the different effects of LDL-C on these two diseases may contribute to discrepancies in the origin of the lipoproteins. LDL-Cs in drusen and Bruch’s membrane are derived from lipids locally produced in the RPE, but not directly deposited from the circulation . AlternaSome potential limitations of the present study should also merit consideration in interpreting the findings. First, although the present results showed that lipids levels may significantly affect the risk of AMD, the results based on cohort studies found that lipids levels only had a tendency for the risk of this disease which was attributed to the fact that the number of cohort studies was relatively limited. Therefore, such associations still need to be investigated to confirm in further well-designed large prospective studies; Second, there is a lack of access to original source data and we are unable to make full use of time-to-event data, therefore the potential bias and confounding effects cannot be completely ruled out; however, the combined sample size was relatively large, and the present results remained robust, adding to the strength of this analysis; Third, even though most of the studies incorporated in the meta-analysis had adjusted for several confounding factors, most studies could not consider genetic susceptibility and lipid lowering medication, and the possibility these confounders might affect the AMD risk interactively with lipid metabolism could not be excluded in the present study. Therefore, further large research studies that allow for the adjustment by more confounding factors, including the genes and environmental factors, especially hypolipidemic drugs and lipid metabolism genes, should be conducted; Fourth, because of the different countries and periods in which the studies were performed, the validated classification and grading systems for AMD were inconsistently applied between studies. Although the stratified analysis showed that different separate criteria of AMD did not alter the direction of the results, such variation may also be likely to lead to an underestimation or overestimation of the association; Finally, the potential publication bias was also a concern. Although we did not observe any apparent publication bias in our statistical tests, it was still difficult to completely rule out this problem.In summary, the results of the present study demonstrated that the participants with high HDL levels may be at increased risk for AMD, whereas higher levels of TC, LDL, or TG had the tendency for a decreased risk of AMD, especially for the early stage of this disease. These results suggest that lipid accumulation may have different pathogenesis in the development of atherosclerosis and AMD. However, the number of cohort studies which have the best epidemiologic evidence was the relatively limited, indicating that further well-designed large prospective studies are needed before definitive conclusions regarding the potential associations between these lipid levels and the risk of AMD can be drawn.


The current findings, however, still remained controversial. Therefore, we performed the meta-analysis to provide a more accurate result. Eligible studies were identified from PubMed database up to June 2015. The association of p27 rs34330 polymorphism and cancer susceptibility was estimated with odds ratios and corresponding 95% confidence intervals. The meta-analysis was performed with Stata 12. A total of ten studies with 11,214 cases and more than 8,776 controls were included in the meta-analysis . In pooled analysis, p27 gene rs34330 polymorphism significantly increased the cancer susceptibility. Subgroup analysis indicated that the elevated risk was observed under all the genetic models for Asians and under three genetic models for Caucasians. Results of sensitivity analysis were similar to the overall results. The results suggested that the p27 rs34330 polymorphism increased the cancer susceptibility, especially in Asians. Further well-designed and large sample size studies are warranted to verify the conclusion.The  Cancer is a leading cause of death and a major public health problem in both economically developed and developing countries. The occurrence of cancer is elevating because of the growth and aging of global population and environmental factors, especially in less developed countries, in which approximately 82% of the world’s population resides. Based on GLBOCAN estimates, about 14.1 million new cancer cases and 8.2 million deaths occurred in 2012 worldwide235678p27/Kip1 (p27) gene  is located on chromosome 12p13 and encodes cyclin-dependent kinase inhibitors (CDKIs) implicated in the negative regulation of the cell cycle1012p27 gene have been identified as associated cancer susceptibility. In 2012, a meta-analysis had been performed to estimate the association between p27 gene rs2066827 polymorphism and cancer susceptibilityp27 gene has also been widely studied for human cancer susceptibility. The existing evidence, however, still remains controversial and has not yet been investigated using meta-analytic methods. Therefore, we aimed in this study to investigate the association of p27 gene rs34330 polymorphism with cancer susceptibility.The 1516171819202122231516171820231921p27 rs34330 polymorphism and cancer susceptibility. Overall, significantly increased cancer susceptibility was observed in all the tested genetic models: homozygous model .Subgroup analysis showed increased cancer susceptibility under all tested genetic models  and no evidence of heterogeneity was found in homozygous model and recessive model. The low to moderate between-study heterogeneity indicated acceptable credibility of the results of this meta-analysis. Moreover, the results of sensitivity analysis were robust when we excluded the studies with controls not in HWE. The results of subgroup analysis according to ethnicity showed that the elevated risk associated with p27 gene rs34330 polymorphism were more predominant in Asians than Caucasians. The results indicated that ethnicity might play an important role in cancer susceptibility; however the underlying mechanism is unclear.It has been postulated that the rs34330 polymorphism, located in the 5′-untranslating region of p27 gene rs2066827 polymorphism was not associated with breast cancer susceptibility. One meta-analysisp27 gene and breast cancer susceptibility, and they found that p27 gene rs34330 polymorphism might be associated with breast cancer susceptibility. In this previous meta-analysisPreviously, two meta-analyses investigated the association of CDKN1B gene polymorphisms (including rs34330 and rs2066827) and CDKN1B rs2066827 polymorphism with susceptibility to breast cancer, respectively30In this study, certain limitations should be taken into consideration. First, confounding factors, such as selection bias and measurement bias, might distort the credibility of the resultp27 gene rs34330 polymorphism might increase the cancer susceptibility, especially in Asians. Further well-designed and large sample size studies are warranted to verify the conclusion.In conclusion, our meta-analysis suggests that the A thorough literature search was performed using PubMed database up to June 2015. We used the following search terms to identify all potential relevant studies: (p27 OR p27Kip1 OR CDKN1B OR “cyclin-dependent kinase inhibitor 1B”) AND (cancer OR carcinoma OR adenocarcinoma OR neoplasm OR tumor) AND . In addition, reference lists of retrieved articles were manually searched. No restriction was applied.p27 rs34330 polymorphism and cancer susceptibility; (3) studies presented detailed genotype counts or odds ratios (ORs) with corresponding 95% confidence intervals (CIs).All the studies included in the meta-analysis should meet the following inclusion criteria: (1) study design was case-control or cohort; (2) studies examined the association between Two reviewers independently extracted the following information from all identified studies according to a standardized data collection form: author name, publication year, ethnicity, location, number of cases and controls, type of cancer, source of control, Hardy-Weinberg equilibrium (HWE) in controls, genotyping method, genetic data, and ORs with corresponding 95% CIs. Any disagreements were resolved by discussion.p27 rs34330 polymorphism and cancer susceptibility. The significance of pooled ORs was determined by Z test at the P < 0.05 level of significance. We estimated the effect of p27 rs34300 polymorphism on cancer susceptibility using homozygous, heterogeneous, dominant, recessive, and allele models. Heterogeneity test was performed with the use of Q statistic at the P < 0.10 level of significance since the chi-squire test has low power in the situation of a meta-analysis when numbers or sample sizes of included studies were small2 metric, a quantitative measurement of heterogeneity among studies2 test. All analyses were performed using Stata 12.0 . The two-sided P values less than 0.05 were considered statistically significant, except where extra specified.The OR and 95% CI were used to measure the association between How to cite this article: Cheng, X.-K. et al. Genetic association between the cyclin-dependent kinase inhibitor gene p27/Kip1 polymorphism (rs34330) and cancer susceptibility: a meta-analysis. Sci. Rep.7, 44871; doi: 10.1038/srep44871 (2017).Publisher's note: Springer Nature remains neutral with regard to jurisdictional claims in published maps and institutional affiliations.