The association between BMI and atrial fibrillation was nonlinear, pnonlinearity < 0.0001, with a stronger association at higher BMI levels, however, increased risk was observed even at a BMI of 22–24 compared to 20. In conclusion, general and abdominal adiposity and higher body fat mass increase the risk of atrial fibrillation.Different adiposity measures have been associated with increased risk of atrial fibrillation, however, results have previously only been summarized for BMI. We therefore conducted a systematic review and meta-analysis of prospective studies to clarify the association between different adiposity measures and risk of atrial fibrillation. PubMed and Embase databases were searched up to October 24th 2016. Summary relative risks (RRs) were calculated using random effects models. Twenty-nine unique prospective studies (32 publications) were included. Twenty-five studies  were included in the analysis of BMI and atrial fibrillation. The summary RR was 1.28  contains supplementary material, which is available to authorized users. The prevalence of overweight and obesity has increased rapidly over the last decades in all areas of the world . OverweiAtrial fibrillation is the most common arrhythmia diagnosed in clinical practice  and globWe searched the PubMed and Embase databases up to October 24th 2016 for eligible studies . A list of the search terms used are provided in Supplementary Tables 1 and 2. We followed standard criteria (MOOSE Guidelines) for reporting meta-analyses . In addiWe included prospective and retrospective cohort studies and nested case–control studies of the association between adiposity measures  and risk of atrial fibrillation that were published in English. Studies in high-risk populations (patient populations), abstract only publications, grey literature and unpublished studies were excluded. Adjusted relative risk (RR) estimates  had to be available with the 95% confidence intervals (95% CIs) in the publication and for the dose–response analysis, a quantitative measure of adiposity and the total number of cases and person-years or non-cases for at least 3 categories of the adiposity variable or on a continuous scale had to be available in the publication. When multiple publications were available from the same study we used the study with the largest number of cases, but when data on different anthropometric measures were covered by different publications from the same study both were included, but each study was only included once in each analysis. A list of the excluded studies and exclusion reasons are found in the Supplementary Table 3.We extracted the following data from each study: The first author’s last name, publication year, country where the study was conducted, study period, sample size, number of cases/controls, exposure variable, exposure level, relative risks and 95% confidence intervals for the highest versus the lowest level of the exposure variable and variables adjusted for in the analysis. Data were extracted by one reviewer (DA) and checked for accuracy by a second reviewer (AS).p < 0.05 was considered statistically significant. If studies reported results separately for men and women or other subgroups we combined the subgroup-specific estimates using a fixed-effects model to generate an overall estimate so that each study was only represented once in the main analysis, but sex-specific results are presented separately in subgroup analyses.We calculated summary RRs and 95% CIs for a 5 unit increment in BMI, 10 cm increment in waist and hip circumference, a 0.1 unit increment in waist-to-hip ratio, and a 5 kg increment in fat mass and weight, 10% increase in body fat percentage, and 5% increase in weight gain using a random effects model . For theThe method described by Greenland and Longnecker  was used2 [p < 0.10. Sensitivity analyses excluding one study at a time were conducted to clarify whether the results were simply due to one large study or a study with an extreme result.Subgroup analyses stratified by sex, measurement versus self-report of adiposity measures, duration of follow-up, geographic location, number of cases, study quality scores, and adjustment for confounders  and potential intermediates  were conducted to investigate potential sources of heterogeneity and heterogeneity between studies was quantitatively assessed by the Q test and I2 . Meta-re2  and Begg2  with theWe identified 29 prospective studies (32 publications) that were included in the systematic review of BMI, waist circumference, hip circumference, waist-to-hip ratio, body fat mass, body fat percentage, weight, and weight gain and risk of atrial fibrillation  compared to women . There was evidence of a nonlinear association between BMI and atrial fibrillation, pnonlinearity < 0.0001  per 5 BMI units.Twenty-five prospective studies (25 publications) , 55–57 i2 = 73%, pheterogeneity = 0.005)  , 29–32 wWaist-to-hip ratio, hip circumference, weight, body fat mass, body fat percentage, pericardial fat, intrathoracic fat, and abdominal visceral fat.2 = 44%, pheterogeneity = 0.15)  , 32, 33 2 = 91%, pheterogeneity < 0.0001; Fig. Three prospective studies (3 publications) , 32, 33 2 = 94%, pheterogeneity < 0.0001)  , 32, 33 2 = 90% pheterogeneity < 0.0001)  , 30, 33 2 = 74%, pheterogeneity < 0.0001)  , 58, 59 2 = 86% pheterogeneity = 0.007)  Fig. b per 5% p = 0.39. When the studies of BMI and atrial fibrillation were stratified by study design, the summary RR was 1.15  for the nested case–control studies and 1.30  for the cohort studies. Study quality was high with a mean (median) score of 7.7 (8) out of 9 points in the analysis of BMI and atrial fibrillation.The positive association between BMI, and risk of atrial fibrillation persisted in almost all subgroup analyses defined by gender, assessment of weight and height, duration of follow-up, geographic location, number of cases, study quality and adjustment for confounding and potential intermediate factors and there was little evidence of heterogeneity between any of these subgroups with meta-regression analyses Table . In furtThis is to our knowledge the first meta-analysis to assess multiple adiposity measures in relation to risk of atrial fibrillation. There was a 28% increase in the relative risk per 5 units increase in BMI, a 18% increase in relative risk per 10 cm increase in waist circumference, a 9% increase in the relative risk per 0.1 unit increase in waist-to-hip ratio, a 32% increase in relative risk per 10 cm increase in hip circumference, a 9% increase in the relative risk per 5 kg increase in body fat mass, and a 10% increase in the relative risk per 5 kg increase in body weight, but no significant association was observed for body fat percentage or weight gain, although the number of studies was very low in these analyses. There was evidence of a nonlinear association between BMI and atrial fibrillation, with a slightly steeper association at higher BMI levels, however, there was evidence of increased risk even within the normal BMI range (22–24) compared to a BMI of around 20, although the increased risk was most pronounced in the obese and severely obese BMI ranges. The association between waist circumference and atrial fibrillation was approximately linear. The positive association between BMI and atrial fibrillation was observed across all geographic locations and both in Caucasians and African Americans, suggesting that adiposity is a risk factor for atrial fibrillation across populations.The current findings are consistent with a previous meta-analysis of 5 cohort studies which found a 39 and 87% increase in the relative risk of atrial fibrillation among overweight and obese, respectively, compared to normal weight subjects . HoweverOur meta-analysis has some limitations that need to be mentioned. Confounding by other risk factors may have influenced the results. However, the association between BMI and atrial fibrillation persisted in subgroup analyses when studies were stratified by whether they adjusted for confounding factors such as age, smoking, alcohol, and physical activity. In addition, the association persisted among studies that adjusted for potential intermediates including hypertension, blood pressure, serum cholesterol, diabetes, coronary heart disease, heart failure, and left ventricular hypertrophy. There was some evidence of heterogeneity between the subgroups of studies that adjusted for heart failure, with weaker, but still significant associations among studies with such adjustment. This could indicate that part of the association between adiposity and atrial fibrillation may be mediated by heart failure. This is consistent with our previous finding of an increased risk of heart failure related to both general and abdominal adiposity  and withMeasurements of weight, height, waist and hip circumferences may have been affected by measurement errors, however, the association for BMI was similar among studies that used measured weight and height compared to those that used self-reported weight and height. Validation studies have reported high correlations between self-reported and measured anthropometric measures –66. BMI Several potential mechanisms could explain an association between body fatness and risk of atrial fibrillation. Adiposity is associated with increased risk of hypertension , insulinOur meta-analysis has several strengths including the prospective design of the included studies which avoids recall bias and reduces the possibility for selection bias, the large number of cohort studies with >83,000 cases and >2.4 million participants in the BMI analysis which provided statistical power to detect moderate associations, the detailed dose–response analyses which clarified the shape of the dose–response relationship, the observation of a similar association between BMI and atrial fibrillation in different geographic regions, and robustness of the findings in multiple subgroup analyses as well as the high study quality of the included studies.The findings have important clinical implications for the prevention of atrial fibrillation as a previous meta-analysis only analysed BMI, but not other fat measures in relation to the risk of atrial fibrillation , and havIn conclusion, our findings confirm that overweight, obesity, abdominal fatness and high body fat mass increase the risk of atrial fibrillation.Supplementary material 1 (PDF 414 kb)Below is the link to the electronic supplementary material.


Intracerebral hemorrhage (ICH) has the highest mortality rate in all strokes. However, controversy still exists concerning the association between plasma homocysteine (Hcy) and ICH. A systematic review and meta-analysis was conducted using Pubmed, Embase, and Web of Science up to April 18, 2017. Standard mean difference (SMD) for mean differences of plasma Hcy levels with 95% confidence intervals (CI) was calculated. Seven studies including 667 ICH patients and 1821 ischemic stroke patients were identified for meta-analysis. Our results showed that Hcy levels in ICH patients were significantly higher than those in healthy controls ; no statistic differences were found in the comparisons of Hcy levels between ICH and ischemic stroke ; further subgroup analysis of ethnicity  and sample size  in relation to Hcy levels between ICH and healthy controls did not change these results. In conclusion, Hcy level may be an aggravating factor in atherosclerosis, which is positively associated with high risk of ICH. Race-specific differences between Asians and Caucasians have no impact on the risk of ICH. ICH is more severe and disabling than ischemic stroke; it has the highest mortality rate of all the strokes4. Though our understanding of ICH has advanced in recent decades, so far this knowledge has not been translated into significant improvement in efficient therapies and prognosis7. Therefore, emphasis has been placed on preventive medicine aiming at reducing traditional risk factors of ICH, mainly hypertension, hyperlipidemia, diabetes mellitus and smoking11. A high level of homocysteine (Hcy) has been regarded as a qualifiable risk factor for ischemic stroke, and Hcy-lowering therapy is widespread used for those patients13. According to guidelines for healthcare professionals from the American Heart Association/American Stroke Association, though no randomized, placebo-controlled study has shown that Hcy-lowering therapy reduces ischemic stroke recurrence, it is recommended that supplements of folic acid, vitamin B6 and vitamin B12 reduce levels of homocysteine and may be considered for patients with ischemic stroke and hyperhomocysteinemia14. However, the relationship between Hcy and ICH remains controversial. Preview study showed no obvious association between Hcy level and risk of hemorrhagic stroke15. Recently, some studies revealed that patients with ICH exhibited higher concentrations of Hcy than that of healthy controls, which implied an underlying mechanism linking plasma Hcy levels with ICH18.Intracerebral hemorrhage (ICH), which accounts for 10–15% of all strokes, is the second most detrimental subtype of stroke20. High level of Hcy is mainly found in patients with a nutritional deficiency of B vitamins as well as being present in genetic diseases where there is an insufficient level of cystathionine beta-synthase enzyme (CβS) or a mutation inn the enzyme 5,10-methylenetetrahydrofolate reductase (MTHFR). Other factors can elevate the level of Hcy i.e. a diet with high methionine content, smoking and a sedentary lifestyle24. According to epidemiological evidence, Hcy is considered to be a potential predictor of pathophysiology of neuronal system disorders26. High level of Hcy induces a neuro-inflammatory process due to its abilities to react with cytokines in the brain, leading to increased permeability of blood-brain barrier (BBB) and impaired integrity of cerebral vessels28. There is convincing evidence that elevated plasma Hcy level is associated with higher mortality rates from ischemic stroke, cardiovascular diseases and peripheral artery disease32. However, it is still controversial whether elevated plasma Hcy level is a risk factor for ICH. Therefore, we performed a systematic overview of retrospective studies to ascertain the association between Hcy and the risk of ICH.Homocysteine, an intermediate sulfhydryl-containing amino acid, is synthesized by the liver in a reaction involving the removal of the methyl group from methionineTo identify eligible studies investigating the association between plasma Hcy levels and ICH, published studies meeting the following criteria were considered eligible for inclusion: (1) case-control studies written in English; (2) studies which assessed the correlation between Hcy and patients with ICH; (3) the cases were confirmed ICH patients and the controls were healthy people; (4) providing information of plasma Hcy levels in the control group and case group at the onset of ICH. The exclusion criteria were as follows: (1) conflicts with the inclusion criteria; (2) publications with duplication or studies with overlapping data from the same study; (3) abstracts, case report, proceedings, letters, reviews, or meta-analysis.Databases of Pubmed, Embase, and Web of Science were comprehensively searched for articles up to April 18, 2017. The terms searched were as follows: (“plasma homocysteine” OR homocysteine OR Hcy OR “plasma Hcy”) AND . We also screened the reference lists of relevant reviews and included relevant articles in these lists.33. A score of up to 9 points was assigned to each study: 4 points for selection, 2 points for comparability, and 3 points for assessment of outcomes. Studies with 6 or more points were considered as high quality, while those with less than 6 were considered to be of suboptimal quality.Two reviewers independently extracted the data and information from the finally selected trials according to a standard protocol, and then discussed or consulted the original report to reach a consensus when discrepancies occurred. In the case of eligible studies, the following information was collected: first author, publication year, country, detection method for plasma Hcy, proportion of men, number of participants as well as plasma Hcy levels in the groups of ICH, control and ischemic stroke. Guidelines of preferred reporting items for systematic reviews and meta-analyses (PRISMA) were followed by the study. Quality assessment was conducted according to the Newcastle-Ottawa Scale (NOS) criteria for retrospective studies2 test  was applied to assess the heterogeneity among studies36. Random effects models were applied for the evidence of significant heterogeneity (I2 test exhibited > 50%); otherwise, the fixed effects model was utilized37. Hcy levels from plasma and serum were used equally. Subgroup analyses by ethnicity and sample size of participants (≥50 or less) were also conducted. Sensitivity analysis was carried out by sequentially omitting one study to examine its influence on the pooled risk estimates. A publication bias was detected by visual inspection of a funnel plot i.e. an asymmetric funnel plot indicated a possible risk of a publication bias.Statistical analysis in this meta-analysis was performed with the software Review Manager 5.2 and STATA version 14.0. Standard mean difference (SMD) with 95% confidence intervals (CI) was calculated by random effects model or fixed effects model to measure the differences in plasma Hcy levels between the case group and control group, the significance of which was assessed by utilizing the Z test. I55. In all, 15 studies were further excluded, including 4 studies with irrelevant outcomes, 3 studies containing ischemic stroke patients, 6 studies in which there was a lack of usable data and 2 prospective studies. Thus, 7 studies were finally included into the current meta-analysis55  whereas the other five studies examined small samples (n < 50). Figure A total of 241 abstracts were retrieved initially. After exclusion of 219 irrelevant or duplicated reports, 22 potentially eligible studies were assessed by reviewing the full text2 = 0%), as well as groups of ICH and ischemic stroke (I2 = 5%); thus, the fixed-effects model was used to pool data. Patients with ICH had a higher serum Hcy level compared with control group   Fig. . There w01) Fig. . The res01) Fig. . A furth01) Fig. .Figure 2Sensitivity analysis performed by omitting one individual study sequentially showed that no single study had any obvious influence on the pooled mean difference Fig. . Since t55. In comparison with a single study, the meta-analysis makes it possible to have a more accurate estimate and evaluation of the impact of Hcy levels on ICH. The findings emerging from comprehensive meta-analysis confirmed that patients with ICH exhibited prominently higher plasma Hcy levels than healthy controls, and this association could not be affected by study quality, participant characteristics and publication bias.As far as we are aware, there has been no relevant systematic review of case-control studies carried out to explore whether plasma Hcy are associated with the risk of ICH. Seven eligible studies with a total of 667 ICH patients were included into our meta-analysis58. In addition, Hcy enhances the release of arachidonic acid from platelets to generate reactive oxygen species (ROS), leading to the deposition of calcium and lipid in the endothelial wall; these changes degrade arterial elasticity and accelerate the process of atherosclerosis61. In the present results, the comparisons of plasma Hcy levels between ICH and ischemic stroke showed no significant difference. This implied that the plasma Hcy level was likely identified as an aggravating factor in atherosclerosis, owing to which the pathophysiology of endothelial degeneration and necrosis of vessel walls could elevate the risk of ischemic stroke and cerebral hemorrhage62.The precise mechanism of Hcy on the susceptibility of ICH remains unresolved. However, our meta-analysis supported that the pathogenesis of ICH likely involves some factor elevating Hcy levels. Experimental studies have verified multiple biological responses triggered by Hcy during the process of atherosclerosis, for example, Hcy inhibits glutathione peroxidase to stimulate the proliferation of smooth muscle cells, by which endothelial nitric oxide synthase (eNOS) can be suppressed via the asymmetric dimethyl-L-arginine (ADMA) pathway, leading to endothelial dysfunction and vascular spasms49. Another study reported that Hcy levels initially decreased within 24 hours and then gradually raised within 48 hours in patients with hemorrhagic stroke, whereas differences in mean Hcy levels at 0, 24 and 48 hours post stroke were insignificant compared with mean Hcy levels in controls54. Accordingly, it is speculated that Hcy levels have been already elevated before the onset of ICH and that their levels could not be affected by acute stress reaction after a stroke. Further studies will be needed to assess the influence of the acute stress reaction during follow-up on the association between Hcy and ICH.However, we are unable to rule out the possibility that the high levels of Hcy may be attributable to the acute stress reaction evoked by stroke, as all blood samples of plasma Hcy were obtained at the onset of ICH rather than before its occurrence. There is one prospective study in which participants were invited to donate a fasting blood sample for future research; this showed that plasma Hcy levels were significantly higher in patients who would develop ICH compared to their healthy controls63. Interestingly, in the present study, the subgroup analysis based on ethnicity in relation to plasma Hcy levels between Asians and Caucasians had no statistic differences, which implied that race-specific differences exerted no effect on the results of our meta-analysis. The reason for this inconsistency might be that race-specific differences between Africans and Caucasians had a profound impact on the risk of ICH rather than race-specific differences between Asians and Caucasians in our study. Although our meta-analysis revealed no statistical heterogeneity, there can also be clinical heterogeneity or methodological heterogeneity between studies. Some possible sources of clinical heterogeneity such as inclusion and exclusion criteria, intervention measures, defining indicators may be the differences. Other sources of methodology heterogeneity such as experimental design, bias risk and outcome integrity may also affect the results.Consistent with the overall results, subgroup analysis on either ethnicity or sample size confirmed that plasma Hcy levels of ICH patients were significantly higher than those of healthy controls. Preview study revealed that Africans had apparently higher risk of ICH compared with Caucasians, suggesting that race played an important role in the ICH risk patternSeveral potential limitations of this meta-analysis must be acknowledged. All included studies followed a retrospective design, which may be more subjected to bias and artifact than prospective studies. The different detection methods employed for plasma Hcy levels may have sensitivity and reliability issues. We were unable to analyze the effect of the acute stress reaction of ICH on plasma Hcy levels since there were insufficient data about Hcy levels before the onset of ICH. Although the result of the funnel plot did not suggest any publication bias, there is the possibility of journals publishing only positive findings as only published data were utilized in the meta-analysis. Additionally, the relatively limited number of studies and the fact that most of patients were Asians, may also influence the sensitivity of statistical analysis. Further randomized controlled trials and larger numbers of participants will be necessary to confirm these findings.Our results of meta-analysis provide support for the hypothesis that elevated level of Hcy in blood is positively associated with the incidence of ICH. Hcy level may be an aggravating factor in atherosclerosis which contributes to ICH. There is no impact of race-specific differences between Asians and Caucasians on the risk of ICH. More clinical trials with more comprehensive data, using larger sample sizes and better design, will be needed to validate our results.


Dear Editors and Authors,b]pyridine (PhIP), 2-amino-3,8-dimethylimidazoquinoxaline (MeIQx) and 2-amino-3,4,8-trimethylimidazoquinoxaline (DiMeIQx) and the “meat-derived mutagenic activity” (MDM).I had the pleasure to read the published article entitled  “DietaryThis work is very necessary and useful for engendering discussion regarding risk factors related to cancer.p < 0.001), MeIQx , DiMeIQx , B(a)P (benzo(a)pyrene)  and MDM . A linear and curvilinear trend was observed in dose–response meta-analysis between CRA risk in association with PhIP, MDM, and MeIQx. “CRC risk  was increased by uptake of MeIQx , DiMeIQx  and MDM . No publication bias could be detected, whereas heterogeneity was in some cases rather high. Mutagenic compounds formed during cooking of meat at high temperature may be responsible of its carcinogenicity” (Abstract).The main findings included “Polled CRA risk  was significantly increased by intake of PhIP (OR (odds ratio) = 1.20; 95% CI : 1.13, 1.28; However, upon careful review of the number of data used to calculate risk in the study, some findings from the meta-analysis appear perplexing. This may result from a substantial problem in repeated usage of data; specifically, possible repeated examinations of OR/RR (relative risk) (95% CI), and missing data in prospective cohort studies. These issues might result in wrong outputs and, thus, skewed findings of the present study.It appears the authors’ above conclusion did not separate the pooled analysis in each study design within the Prospective Cohort Study and Case-Control Study. Those designs definitely differed regarding exposure measures of HCA intake before the occurrence of colorectal adenoma and cancer (Prospective Cohort Study) and after these diseases occurred (Case-Control Study). The findings presented  and controls  from the North Carolina Colon Cancer Study ,10 from Were the findings of the meta-analysis modified due to repeatedly inputting data twice, three or four times? The authors should correct and fix these problems to show accurate results. For the study by Sinha, et al. in 2005 , the autFor the study by Rohrman, et al. in 2009 , the autFor the study by Ferrucci, et al. in 2012 , the autBy repeated usage of RR (95% CI), there were four published articles of prospective cohort studies only; the number of data points used to calculate the risk were eight for PhIP, six for MeIQx, six for DiMeIQx, and five for MDM .For the two studies by Miller, et al. in 2013  and JoshFor the study by Cross, et al. in 2010 , the autAfter the doubled RR(95% CI), there were only three published articles of prospective cohort studies, and the number of data points used to calculate the risk were five for PhIP, MeIQx, and DiMeIQx, and four for MDM .Among the available three prospective cohort studies, only the study by Cross, et al. in 2010  showed aAgain, were the findings of the meta-analysis modified due to double input of OR/RR (95% CI) data? The authors should correct and fix these problems to show accurate results.p Value = 0.005 that might not reflect the true findings of the pooled analysis of only two available studies of prospective cohort studies. The data of MDM was missing from the study by Ollberding, et al. in 2012 . The poiI address these points only to illustrate the subtleties in handling data used to calculate risk and the importance of avoiding repeated usage of cases and controls due to multiple publications. For example, for OR/RR (95% CI), if analysis includes CRC, then the usage of their sub sites of colon and rectum in the computer analyzing programs will result in errant data in a meta-analysis. The authors should rerun the meta-analysis after excluding repeated cases and controls and avoid double OR/RR (95% CI). I believe this will result in accurate outputs and findings.


When the adjusted data were analyzed, the summary estimate decreased to 25% reduction in colorectal adenomas risk . Besides, a significant reduction of colorectal cancer risk was also observed . And when the adjusted data were analyzed, colorectal cancer risk for metformin users was decreased with a reduction of 22%, compared with non-metformin users and other treatment users . Our meta-analysis suggested that metformin therapy may be associated with a decreased risk of colorectal adenomas and colorectal cancer in type 2 diabetes mellitus patients.Recent evidence indicates that metformin therapy may be associated with a decreased colorectal adenoma/colorectal cancer risk in type 2 diabetes patients. However, results are not consistent. We therefore performed a systematic review and meta-analysis to assess the association between metformin therapy and risk of colorectal adenomas/colorectal cancer in type 2 diabetes mellitus patients. We searched the literature published before Aug 31, 2016 in four databases: PubMed, Embase database, CNKI and VIP Library of Chinese Journal. Summary risk estimates (adjusted OR/adjusted RR/adjusted HR) with their 95% confidence interval (95% CI) were obtained using a random effects model. Twenty studies  were selected in terms of data of colorectal adenomas or colorectal cancer incidence. Metformin therapy was found to be associated with a decreased incidence of colorectal adenomas (unadjusted OR=0.80, 95% CI: 0.71-0.90,  According to WHO, over 220 million people are suffering from diabetes disease. The International Diabetes Federation estimates that the number of people with impaired glucose tolerance will increase to 472 million (9 % of the adult population) by 2025, and around 40-50% of them will develop type 2 diabetes mellitus (T2DM) . Many stMetformin is a potent anti-hyperglycemic agent. It can reduce hyperinsulinemia, improve insulin resistance , and lowp=0.0115). Similar results were also reported in other studies [Some researchers indicated that metformin may be a potential protective factor of colorectal adenomas and colorectal cancer in T2DM patients. Kim et al  and Cho  studies , 13. How studies , 14, 15, studies , 17, andTo explore associations between metformin therapy and colorectal cancer and adenomas risk, we identified available studies to make a quantitative meta-analysis for the purpose of more cognition of using metformin in T2DM patients. This investigation is important to identify the existing knowledge on the effects of metformin therapy in T2DM patients.We identified 759 unique papers after searching PubMed, Embase database, China National Knowledge Infrastructure (CNKI), and VIP Library of Chinese Journal. Based on the inclusion and exclusion criteria defined above, we excluded 739 articles. A total of 20 full-text articles were included in this meta-analysis Figure . Among tFive studies, including case-control studies , 23, 27 p=0.0002). Low between-study heterogeneity was found   and retrospective cohort studies   .p=0.03), low between-study heterogeneity was found   .For we only have five studies refer to metformin therapy and colorectal adenomas risk, the number is too small to conduct the Begg's and Egger's test. However, the funnel plot illustrated a symmetrical distribution of the points, suggesting a lack of publication bias .Fifteen studies  for metformin users (metformin exposure more than 1 year) compared with non-metformin users or other treatment users. High between-study heterogeneity was found  , the between-study heterogeneity was decreased   Figure .p=0.70). However, a significant reduction of colorectal cancer risk was observed in retrospective cohort studies   Figure .p=0.0001). High between-study heterogeneity was found  (p=0.13). Moderate between-study heterogeneity was found  (p=0.002). Moderate between-study heterogeneity was found  , a significant reduction of colorectal cancer risk was observed in adjusted OR data  . After o p=0.14) . And a sp=0.001) p=0.127) (p=0.182)   and Eggep=0.182) . The funp=0.0002) and colorectal cancer  among T2DM patients, compared with non-metformin users and other diabetic drug users. The degree of heterogeneity among studies may cause by the difference in study design, populations or comparators. However, almost all of the results showed a protective effect of metformin.This meta-analysis suggests that metformin therapy may be associated with a decreased risk of colorectal adenomas . Similarly, Monica Franciosi et al [p=0.0009). However, Andrea DeCensi et al [Our results are consistent with previous observations suggesting a protective role of metformin for colorectal cancer , 17, 29.si et al  found ansi et al , which wsi et al .p<0.001). Zhu et al [p=0.019). And a study by Tseng et al [p<0.0001). These studies confirmed that metformin therapy may serve as a protective factor in decreasing cancer risk.Metformin also showed a protective effect in many other cancers. A meta-analysis conducted by Wu et al  showed thu et al  confirmeng et al  proved mEvidences from previous studies indicate that metformin may interfere with carcinogenesis through direct (insulin-independent) and indirect (insulin-dependent) mechanisms. Metabolic syndromes in T2DM patients such as insulin resistance, hyperinsulinemia and hyperglycemia are risk factors of several cancers , 36. MetCompared with other existing studies, our meta-analysis has several advantages. Firstly, our analysis included a much larger number of studies, and provides more precise results. Sensitivity analysis also indicated that our pooled risk estimates were considerably credible. Secondly, we adopted more rigorous criteria for study selection to ensure the reliability of the meta-analysis, taking into account factors which may affect the results. In addition, we analyzed the results separately by type of diseases , which provided a stronger evidence of protective effect of metformin among T2DM patients.However, our study also has several limitations. Firstly, the selection of populations is inconsistent in different studies, which may be an important source of heterogeneity. Secondly, the description of drug dose and length is unclear among studies. Most studies usually classify metformin as two groups: metformin users vs. non-users or other treatment user. And few studies collected metformin dose data. There might have a dose–response relationship between metformin use and cancer incidence for T2DM patients, and we can't calculate it, which may be another large source of high heterogeneity in our studies. Thirdly, different adjustments may influence the summary estimation of the association between metformin treatment and colorectal adenomas/colorectal cancer risk. Furthermore, the protective effect of metformin may be overestimated due to the association between insulin using and higher risk of colorectal cancer . Future In conclusion, our meta-analysis indicates that metformin is associated with a reduction in risk of colorectal adenoma and colorectal cancer incidence in individuals with T2DM, compared with other diabetic treatments. Our results also support the hypothesis that in T2DM patients with high risk for colorectal adenomas/colorectal cancer (due to their metabolic condition), the candidates for drug therapy may better be treated with metformin. And further prospective studies, especially large-scale clinical studies and cell experiments are needed to confirm the relationship between colorectal cancers and metformin treatment.Our meta-analysis was in accordance with the Systematic Reviews and Meta-Analysis guidelines: PRISMA Checklist.A comprehensive literature search was performed in the PubMed, Embase database, China National Knowledge Infrastructure (CNKI) and VIP Library of Chinese Journal from their earliest available data to Aug 31, 2016. The following search terms and combinations were used in keyword and subject heading search: (“diabetes” or “diabetes mellitus” or “DM” or “insulin resistance“) and (“therapy” or “treatment” or “therapeutics” or “therapeutics” or “metformin” or “biguanides“) and  or (“neoplasm” or “neoplasia” or “cancer” or “tumor” or “carcinoma” or “adenoma“)). Only human studies were included. The languages of included articles were limited to English and Chinese. A manual search was also conducted for references cited in the selected articles. (refer supplementary for detailed search strategy).We applied five inclusion criteria on retrieved articles. The eligible paper should: (1) be an original article; (2) report randomized or observational studies; (3) study patients with T2DM ; (4) have metformin therapy as the first treatment (metformin exposure more than 1 year) compared with non-metformin user (never use metformin) and other treatments ; (5) provide relative risk (RR), odds ratio (OR) or hazard ratio (HR) with their 95% confidence interval (95% CI), or provide sufficient data to allow adequate estimation of the RR/OR and 95% CI. We dropped the paper if it: (1) cases report, summary, review or meta-analysis; (2) had no individual data on type of diabetes; (3) included patients with type 1 diabetes; (4) researched subjects with complication (diabetic retinopathy etc.), hereditary colorectal syndromes, history of colorectal cancer, chronic inflammatory bowel disease, or previous bowel resection; (5) reported duplicated results or unavailable data.Two authors (FFL and HX) selected studies independently according to the inclusion and exclusion criteria listed above. The characteristics extracted were the name of first author, publication year, study design, race, study period, adjust factor or covariance, crude numbers of exposed and unexposed subjects, crude numbers of colorectal adenomas/colorectal cancer in exposed group and unexposed group (for cohort studies), or crude numbers of cases and controls, crude numbers of colorectal adenomas/colorectal cancer in case and control group (for case-control studies), RR, OR or HR with their 95% CI. When there were multiple publications from the same population, only the study with largest sample size were included. Conflicts of data extraction between the two reviewers were resolved by discussion and consensus with an arbitrator (SFN).Two authors (DSR and YSL) conducted quality assessment of included observational studies (cohort study and case-control study) independently according to The Newcastle-Ottawa Scale (NOS). This sc2 statistics [2 >75.0%, 50.0–75.0% and I2 <50% indicate low, moderate, and high heterogeneity respectively. Sensitivity analysis was conducted to identify studies contributing disproportionately to the observed heterogeneity, by omitting a study at a time to analyze the influence of individual studies on the summary estimate [p value less than 0.05 was considered statistically significant. RevMan (Version 5.2) was used for these calculations.In this meta-analysis, we used a random-effects model to estimate the pooled effect across studies for the association between metformin therapy and the risk of colorectal adenomas/colorectal cancer. For the random-effects model accounts for variations between studies in addition to sampling error within studies . We calcatistics . I2 >75.estimate . The sigp value was less than 0.05. These analyses were carried out with Stata (Version 12.0).The possibility of publication bias, which may be caused by the non-publication of small number of studies with negative findings, was assessed with a funnel plot for asymmetry. Besides, the Begg's test and Egger's test were also conducted to assess the publication bias . The met


A total of 17 studies  were included. When compared with subjects having BMI < 25, individuals with BMI 25–30 had significantly higher risk of CRA . Subjects with BMI ≥ 30 had similarly higher risk of CRA . The heterogeneity was mild to moderate among studies. The associations were significantly higher than estimates by previous meta-analyses. There was no publication bias detected . Subgroup analysis showed that the magnitude of association was significantly higher in female than male subjects  being insignificant in Africans; and no difference exists among different study designs. In summary, the risk conferred by BMI for CRA was significantly higher than that reported previously. These findings bear implications in CRA risk estimation.The objective of this meta-analysis is to evaluate the odds of colorectal adenoma (CRA) in colorectal cancer screening participants with different body mass index (BMI) levels, and examine if this association was different according to gender and ethnicity. The EMBASE and MEDLINE were searched to enroll high quality observational studies that examined the association between investigator-measured BMI and colonoscopy-diagnosed CRA. Data were independently extracted by two reviewers. A random-effects meta-analysis was conducted to estimate the summary odds ratio (SOR) for the association between BMI and CRA. The Cochran’s Q statistic and IThe online version of this article (10.1007/s10654-017-0336-x) contains supplementary material, which is available to authorized users. Its inct of CRC –7. Whilst of CRC . Its incColorectal adenomas (CRA) are present in more than 30% of asymptomatic general populations . Among aNevertheless, the association between BMI and CRA has not been consistently demonstrated in all populations –28. Somewww.greylit.org), related thesis, and conference reports. No language restrictions were imposed.We conducted the literature search by systematically searching MEDLINE (from 1946 to March 2017), EMBASE (from 1974 to March 2017) and by hand searching the reference lists of original studies and review articles on this topic. Our search terms consisted of three main components, colorectal  AND disease (cancer* OR neoplas* OR tumor* OR tumour* OR carcinoma* OR sarcoma* OR adenoma* OR lesion* OR polyp* OR CRC) AND obesity or overweight    with 95% confidence intervals (CI) between CRA and BMI categories were recorded. We excluded the following studies: (1). those with hyperplastic polyps, serrated adenomas or CRC cases as majority of all lesions; (2). those where subjects had higher CRC risk as compared to the general population, for instance, individuals with a family history of CRC in first-degree relatives; (3). those that could not generate OR for BMI category and CRAs; (4). those with symptomatic participants; (5). those with BMI data obtained from self-reported questionnaires; (6). those with CRA not diagnosed by colonoscopy and histological examination; (7). those with CRA data not derived from the whole colon and rectum. The eligibility of studies was assessed by two investigators (J. L. H. and C. H. C.) and in cases of disagreement, consensus was made via referral to a third reviewer (M. C. S. W.). We attempted to contact authors of studies if there were any missing data.The Newcastle Ottawa Scale (NOS) was employed to evaluate the quality of the included studies according to their design by two assessors (J. L. H. and C. H. C.) who are librarian experts , 37. TheFor cross-sectional studies, additional items for quality assessment include: (1) sample size: if the sample size is justified, one point was assigned, otherwise no point was given; (2) non-respondents: one point was assigned if the response rate is satisfactory, otherwise no point; (3) assessment of the outcome: for those studies in which outcome assessment was independent and blinded, one extra point was added accordingly; (4) statistical test: one point was assigned if the statistical test is appropriate, clearly described and complete; otherwise no point was assigned. For case-control studies, additional items include: (1) same method of ascertainment for cases and controls: one point was assigned; (2) non-response rate: one point was given if the rate for both case and control groups was the same, and no point was assigned for non-respondents. For cohort studies, the additional items are: (1) demonstration that outcome of interest was not present at study commencement: one point was assigned for stating exclusion of CRA/ advanced CRA/ CRC subjects or stating subjects have no history of CRA/ advanced CRA/ CRC; (2) follow-up duration: one point was assigned for all eligible studies if the follow-up period is long enough to detect CRA; (3) adequacy of following up of cohorts: one point was assigned for completing at least 90% of follow-up. Scores ranged from 0 (lowest) to 9 (highest). Similar to previous literature , studies2), overweight (25–30 kg/m2) and obese (≥ 30 kg/m2). The number of cases and non-cases in the 3 categories, as well as the study design, gender and subject ethnicity, were recorded if available. Data extraction and data checking were performed by 3 investigators The characteristics of studies were recorded, including the names of the first authors, publication year, country, design, enrolment year, BMI category, strategies to capture BMI data, the definition of non-cases and the definition of advanced adenoma. BMI is categorized according to WHO classification: normal (< 25 kg/m2) with BMI < 25 kg/m2. The SOR was computed with the assumption that the outcomes categorized by different BMI groups were derived from patients independently, so there was no within-study correlation of adenoma prevalence. Z-tests were used to investigate the significance of pooled estimate, and Cochran’s Q and I2 statistics were used to examine the heterogeneity within groups and between groups [Random effects model meta-analysis was conducted to synthesize a summary estimate of the association between different BMI groups and CRA. Summary odds ratios (SOR) with 95% confidence intervals (CI) were used as a proxy measure for effect size, and were calculated by comparing 3 BMI categories  with metafor package ver. 1.9–9 was used to conduct the statistical analysis . All funuideline .2 was 29.3, 49.7 and 58.1% in Asian, African and white subjects, respectively. No studies were found to use identical cohorts. The search did not identify any studies published in grey literature.The search strategy yielded 3292 citations. We removed 1027 duplicates, and 2173 articles were removed after title and abstract review   and did not show statistically significant difference when compared with any CRA. When compared with subjects with BMI < 25, the odds of advanced adenoma was significantly higher among those with BMI ≥ 25 . The relationship between BMI and advanced adenoma using “non-advanced adenoma” as non-cases did not show statistical significance.Meta-analysis of the included articles via a random-effects model showed a SOR of 1.42  among subjects with BMI ≥ 25 compared to subjects with BMI < 25, where the heterogeneity was moderate and statistically insignificant (Ip difference of < 0.001) . Meta-regression analysis based on BMI 25–30 as a reference and BMI>30 kg/m2 implied that different levels of BMI could not explain the heterogeneity observed in this meta-analysis .Eight studies examined the association in men and women separately, and it was found that female subjects had significantly higher odds of CRA  when compared with men  and Begg’s test  for funnel plot asymmetry identified insignificant publication bias  is associated with similar risk for CRA when compared with obesity (BMI ≥ 30), and hence bring forth an alert to physicians and public health practitioners on early intervention for overweight patients in order to reduce future risk of CRA. In addition, our data showed that risk algorithms for CRA would need to take gender and ethnicities into account for more accurate risk prediction, and these findings could be used for devising such risk-stratification scores. Future studies should examine the mechanistic aspects of the differential effects of these variables on CRA development. As there is a scarcity of prospective studies on the impact of BMI on progression of CRA to advanced CRA, additional longitudinal cohort evaluations should be performed with strategies that address confounding and selection biases.Supplementary material 1 (DOCX 16 kb)Supplementary material 2 (DOCX 40 kb)Below is the link to the electronic supplementary material.


PCSK9 rs505151 and rs11591147 polymorphisms are identified as gain- and loss-of-function mutations, respectively. The effects of these polymorphisms on serum lipid levels and cardiovascular risk remain to be elucidated.PCSK9 rs505151 and rs11591147 polymorphisms with serum lipid levels and cardiovascular risk by calculating the standardized mean difference (SMD) and odds ratios (OR) with 95% confidence intervals (CI).In this meta-analysis, we explored the association of PCSK9 rs505151 G allele was related to higher levels of triglycerides  and low-density lipoproteins cholesterol (LDL-C)  and increased cardiovascular risk . The rs11591147 T allele was significantly associated with lower levels of total cholesterol (TC) and LDL-C  and decreased cardiovascular risk  in Caucasians.Pooled results analyzed under a dominant genetic model indicated that the PCSK9 rs505151 confers increased triglyceride (TG) and LDL-C levels, as well as increased cardiovascular risk. Conversely, the variant T allele of rs11591147 protects carriers from cardiovascular disease susceptibility and lower TC and LDL-C levels in Caucasians. These findings provide useful information for researchers interested in the fields of PCSK9 genetics and cardiovascular risk prediction not only for designing future studies, but also for clinical and public health applications.This study indicates that the variant G allele of  Cardiovascular disease (CVD) is the leading cause of death and contributes substantially to the heavy disease burden worldwide . It is aPCSK9 gene that cause an autosomal dominant form of hypercholesterolemia (ADH). Extensive research into the function of PCSK9 has since been conducted. To date, the best characterized property of PCSK9 is its ability to enhance the intracellular degradation of the LDL receptor (LDLR), which mediates approximately 70% of LDL-C clearance. Secreted PCSK9 in the circulation effectively binds to the LDLR on the surface of hepatocytes, thereby targeting the LDLR for lysosomal degradation and preventing recycling to the hepatocyte surface, thus leading to considerable elevation in LDL-C levels [Proprotein convertase subtilisin/kexin type 9 (PCSK9), which was identified as the ninth member of the proprotein convertase family in 2003, plays a key role in lipid metabolism, and has emerged as an important modulator of cardiovascular health [C levels .PCSK9 gene is located on the small arm of chromosome 1p32.3 and comprises 12 exons and 11 introns [PCSK9 rs11591147  variant contains a replacement mutation (arginine to leucine at position 46) located in exon 1. This relatively rare variant is considered to be a LOF-mutation of PCSK9 [The  introns . It is h introns . These f introns . PCSK9 r introns . The PCSThe current meta-analysis was performed according to the principles proposed by the Human Genome Epidemiology Network (HuGeNet) HuGE Review Handbook of Genetic Association Studies , 11.Studies dealing with the associations of the two SNPs (rs505151 and rs11591147) with plasma lipids levels and risk of cardiovascular disease in humans were considered eligible. Relevant studies were searched in PubMed, Chinese National Knowledge Infrastructure and WANFANG database. The search work was last updated on September 1, 2016. The following three groups of keywords we performed by searching MEDLINE (via the PubMed gateway): “proprotein convertase subtilisin/kexin type 9” OR PCSK9 OR “neural apoptosis-regulated convertase 1” OR NARC1, polymorphism OR SNP OR “single nucleotide polymorphism” OR variant OR variation OR mutation, lipid OR dyslipidemia OR “coronary heart disease” OR “myocardial infarction” OR “coronary artery disease” OR “ischemic heart disease” OR “acute coronary syndrome” OR “CAD” OR “CHD”. References from the retrieved articles and previous meta-analysis were searched manually for additional qualified studies.The studies eligible for the meta-analysis must meet all the following inclusion criteria: (i) case-control or cohort studies; (ii) contained rs505151 and/or rs11591147 genotype data; (iii) adequate data for calculating the standardized mean difference (SMD) and odds ratios (ORs) and correspond 95% confidence intervals (CIs). Exclusion criteria were as follows: (i) studies did not provide sufficient data to extract the information we needed; (ii) case report, review, meta-analysis, cell line and animal experiment studies; (iii) repeated publication about the same population.Two investigators (Qiu and Li) screened all the records and extracted data independently, the third investigator (Zhang) was involved in discussing to avoid bias when there were disagreements between Qiu and Li. Following information was extracted from each of the eligible studies: first author, year of publication, ethnic groups of the patients, type of study, sample size, genotyping method, age, sex, minor allelic frequency (MAF); Hardy-Weinberg equilibrium (HWE).PCSK9 rs505151 and rs11591147 genotype distributions were assessed by Fisher’s exact test. A p < 0.05 for the test was considered deviated from the HWE. The pooled SMD with 95% confidence interval (CI) was applied to calculate the differences of plasma lipid levels between different genotypes. The OR and corresponding 95% CI were used to evaluate the strength of the association between the polymorphisms of two SNPs and cardiovascular risk. Dominant genetic model was conducted to assess the genetic associations, the reasons for the choice are as follows: (i) PCSK9 rs505151 and rs11591147 polymorphisms are rare in human and low MAF were presented in candidate gene studies, on the premise that the difference between carrying one and two copies of the genetic variant is likely to have less effect on the effect size (OR and SMD), perhaps a dominate mode is most reasonable. (ii) For some studies, only dominant genetic data were available; (iii) one model does not require adjustment for multiple hypotheses (which is necessary when different models are used); however, dominant model is commonly used in genetic association synopses.The deviations from the HWE for the 2 statistics [p < 0.10 for the Q test was considered statistically significant. For I2, which describes the proportion variation in point estimates that is due to variance rather than within-study error, the value of I2 ranged from 0 to100% indicates different degree of heterogeneity . Meta-regression, meta-sensitivity and subgroup analysis were conducted to explore the sources of heterogeneity when p < 0.10 for the Q test. SMD, ORs and corresponding 95%CI were calculated by performing fixed effect meta-analysis when the heterogeneity was under the moderate degree or not exist; in otherwise, the analysis model reduced to a random effect meta-analysis. The choice of this model was suggested mainly by the heterogeneity mostly expected in genetic association studies. Meanwhile, the potential bias was assessed by statistical evaluation with Begg’s rank correlation [Between-study heterogeneity was assessed by the chi-square-based Q test and Iatistics . A p < 0relation  and Eggerelation  while thThe α level of significance was set at 0.05, except for the Q-test (0.10).All statistical analyses were performed with with STATA/SE.12.0 , Review Manager Version 5.3 .Amount of evidence, mainly based on the study sample size, was graded by the sum of subjects carrying the variant allele , category “A” corresponds to a sample size over 1000, “B” and “C” correspond to 100–1000 and <100, respectively.2, where I2 < 25% was considered as “A”, 25–50% and >50% were identified as “B” and “C”, respectively.Replication of genetic associations were depended upon the between-study inconsistency defined by IProtection from bias was graded as “A” if there was no notable bias or may not affect the presence of the association; in category “B”, bias could be present or there was considerable missing information on the generation of evidence; in category “C”, demonstrable bias that can affect the presence or absence of the association.The Venice criteria was applied to assess the credibility of each nominally statistically significant association identified by meta-analysis . Three cFollowed the three letters stated above, evidence was categorized as “strong” (A grades only), “weak” (one or more C grades) or “moderate” .The quality of summary evidence for no association was also assessed in the meta-analysis. We calculated the power instead of the sample size; the other aspects including replication and protection from bias were also accounted for according to the Venice criteria . The powCKS9 rs505151 polymorphism, 15 articles comprising 14,451 subjects were identified from the initial search corresponded to the plasma lipid levels, 3373 (23%) were Asians and 11,078 (77%) were Caucasians. The MAF ranged greatly from 2.1%to 14.7%. Genotype distributions in four studies deviated from HWE and two studies did not provide sufficient data to evaluate genotype distributions. There were 12 case-control articles including 11,203 subjects related to the association between rs505151 and cardiovascular disease risk. The constituent ratios of Asians and Caucasians were 23% and 77% respectively. In this group, the MAF varied from 3.8% to 7.5%, and it lower than the reported frequency (10%). Among them, only one study of the genotypes distribution in controls deviated from HWE, and one study was failed to assess the genotype distributions. We identified 8 eligible articles encompassing 17,090 subjects to study the association between PCSK9 rs11591147 variation and plasma lipid levels, most cases were Caucasians (98%), MAF of this group higher than the reported frequency (0.6%) and it varied from 1.6% to 25.3%. Except four articles that could not to be assessed the distribution of genotypes, it did not deviated from HWE in the rest 6 articles. Four case-control studies and three cohort studies revealed the association between PCSK9 rs11591147 polymorphism and cardiovascular disease risk, a total of 60,677 subjects included in this group and all of them were Caucasians. The MAF ranged from 1% to 1.7%. The distribution of genotypes in controls of these studies did not deviated from HWE, except one study that could not to be evaluated. The details of characteristics of each individual study were demonstrated in Table A total of 32 studies met the inclusion criteria and were included in the final meta-analysis, the process of study selection were shown in Fig. PCSK9 rs505151, pooled results under dominant genetic model (AG + GG VS AA) indicated that G allele carriers had higher TG levels in Asians  and higher LDL-C levels  and systolic blood pressure (SBP) Table As for P = 0.000, I2 = 0; LDL-C, SMD: -0.44, 95% CI: -0.55 to −0.33, P = 0.000, I2 = 0, Fig. Findings showed that T allele in rs11591147 was significantly associated with lower TC level and LDL-C level in Caucasians , the evidence was identified as “moderate”; subgroup meta-analysis, which stratified by ethnicity presented consistent results . Conversely, the T allele of rs11591147 related to decreased cardiovascular risk in Caucasians .As shown in Fig. PCSK9 rs505151 polymorphism with increased serum TG concentration and increased cardiovascular risk, the associations between PCSK9 rs11591147 polymorphism and decreased cardiovascular risk were based on heterogeneous data, which lowered the credibility of pooled evidence. Begg’s and Egger’s tests were applied to detect the potential publication bias, data showed there was no potential publication bias in all the comparisons except one meta-analysis about the association of rs11591147 polymorphism and plasma HDL-C level (P = 0.026 for Egger’s tests). More details were reported in Table Some heterogeneity was observed in the meta-analysis. Among these statistical Significant findings, the associations of PCSK9 with serum lipids level and cardiovascular risk that has been conducted to date. Synthetic results clearly showed an association between the G allele of PCSK9 rs505151 and increased serum LDL-C levels; this is the first time that the relationship between PCSK9 rs505151 variants and increased TG concentrations has been demonstrated. Furthermore, this single nucleotide polymorphism (SNP) was also shown to be related to an increased incidence of cardiovascular events. Conversely, the T allele of the PCSK9 rs11591147 variation was found to be associated with reduced serum TC and LDL-C levels and strongly related to a reduction in cardiovascular risk among the general Caucasian population. According to the Venice criteria [In the current study, we performed the most comprehensive meta-analysis of the associations of the rs505151 and rs11591147 functional mutations of PCSK9 rs505151 variants with increased serum TC and LDL-C levels, as well as increased cardiovascular risk [PCSK9 rs505151 variants and serum TG levels, although it remains to be determined whether this relationship is causal or a concomitant phenomenon. As for PCSK9 rs11591147, a meta-analysis has been reported that the PCSK9 46 L allele was associated with reductions in LDL-C and ischemic heart disease via pooled three independent studies in 2010 [PCSK9 rs11591147 variants with lipid levels and cardiovascular risk; the consistency of the results of this meta-analysis further confirmed the robust relationship.Previous meta-analyses have shown the association of lar risk –20; howe in 2010 . Given tPCSK9 are classified as GOF- and LOF-mutations, respectively. Notably, this study further confirmed the association of rs505151 with increased LDL-C levels and cardiovascular risk, while there was a strong association of the rs11591147 polymorphism with reduced LDL-C levels and cardiovascular risk. The PCSK9 gene is highly polymorphic, and functional variants affect the activity of the PCSK9 protein, resulting in lipid metabolism disturbances. Despite the less marked effect of a single SNP on the pathophysiological processing, adding genetic information to lipid management and cardiovascular risk prediction may be potentially useful in clinical practice. Pharmacogenetic studies have shown that GOF and LOF variants of PCSK9 are associated with worse and better responses to statin therapy, respectively [PCSK9 rs505151, into risk prediction models, may show some improvement in cardiovascular risk prediction for primary prevention. Unlike the traditional factors, genetic variants, if they can be identified, may be strong predictors with lifelong value in preventive management.Since it was first reported in 2003, PCSK9 has attracted a lot of attention regarding its key role in lipid metabolism. PCSK9 enhances LDLR lysosomal degradation, resulting in reduced LDL-C clearance, thereby leading elevated serum LDL-C levels , 23. Funectively , 25. ThePCSK9 rs11591147 was small, and pooled results revealed heterogeneity in the Asian group, but not in the Caucasian group, indicating that more high-quality clinical studies in Asian populations are required. Fourth, most original studies included in the present meta-analysis used the combined cardiovascular event to estimate the cardiovascular risk, thus taking difficult to identify the risk of specific cardiovascular event. In the final, though we have provided evidence support the association between the two SNPs of PCSK9 (rs505151 and rs11591147) with lipid traits and cardiovascular events, we compromised estimates of heritability based on the current data. Accurate estimates of heritability will require more extensive examination of each identified SNP, especially in a scenario where variants are more likely to be causal for traits and disease.Some limitations of this meta-analysis should be noted. First, heterogeneity in the data may reduce the credibility of the pooled evidence. The main factors responsible for this heterogeneity were study design (cohort or case-control) and genotyping method (Taq Man or PCR-RFLP), which is a common problem in genetic meta-analyses. Therefore, the adoption of strict standards will be encouraged in performing clinical studies. Second, only one genetic model (dominant model) was applied in our analysis, as explained in the Materials and methods section. However, the use of different models would have increased the number of meta-analyses, with consequent inflation of the type I error . Third, PCSK9 rs505151 allele confers increased TG and LDL-C levels on the carrier, as well as increased cardiovascular risk. Conversely, the variant rs11591147 allele protects against CVD susceptibility and is associated with lower TC and LDL-C levels. These findings could provide useful information for researchers interested in PCSK9 and cardiovascular risk prediction not only in the design of future studies, but also improved clinical and public health. However, further investigations are required to identify the biological function of the two PCSK9 SNPs and to distinguish direct or indirect influences of the variant alleles on cardiovascular risk.This study provides evidence that the variant


Objectives. Apo E genes involved in lipoprotein synthesis and metabolism are considered one of the candidates to CHD. However, the results remain conflicting. Methods. We performed this meta-analysis based on 30 published studies including 11,804 CHD patients and 17,713 controls. Results. Compared with the wild genotype E3/3, the variant genotypes ApoEE3/4 and E4/4 were associated with 22% and 45% increased risk of CHD, respectively . Besides, compared with ε3 allele, carriers with the ε4 allele had a 46% increased risk of CHD , while the ε2 had no significantly decreased risk of CHD. In the subgroup analysis by ethnicity, ε4 had a 25% increased risk of CHD in Caucasians , and the effects were more evident in Mongolians . The ε2 allele had a decreased risk of CHD in Caucasians , but not in Mongolians. Conclusions. The analysis suggested that ApoEε4 mutation was associated with the increased risk of CHD, while ApoEε2 allele had a decreased risk of CHD just in Caucasians. Coronary heart disease (CHD) as a multifactorial disease caused by genetic and environmental factors became one of the leading causes of mortality and morbidity worldwide, especially in the developed countries . Previou Apo E phenotype among 45 different populations in a meta-analysis [ε4 allele were associated with 26% increased risk of CHD compared with ε3 allele [ Apo E genotypes on CHD risk in the general population. Therefore, the present meta-analysis is designed to derive a more plausible estimation.In 1992, Dallongeville et al. tested the consistent relationship between plasma TG levels andanalysis . Another3 allele . Since t3 allele –17. Howe Apo E and polymorphism(s) single nucleotide polymorphism (SNP), variant, mutation and coronary artery disease, coronary heart disease, CAD, and CHD. Hand searches were also performed.We search the electronic databases PubMed, Web of Science, Embase, Wanfang, China National Knowledge Internet (CNKI), and VIP for relevant English language articles from Jan 1, 2000, to Mar 1, 2016, using the following index terms: Apo E gene polymorphism in CAD or CHD; (2) case-control studies; (3) studies with full-text articles; (4) sufficient data for calculating an odds ratio (OR) with 95% confidence interval (CI); (5) not republished data.The inclusion criteria were as follows: (1)ε2, ε3, and ε4, and genotype distribution in case-patients and controls. Definition of CHD end points includes MI, angina pectoris, percutaneous transluminal coronary angioplasty, coronary artery bypass graft surgery, and severe stenosis on coronary angiography. In most studies that included case-patients with coronary artery disease, diagnosis was based on angiographically documented luminal stenosis (≥50%) in at least 1 of the 3 major coronary arteries.All the data were carefully extracted from all eligible publications independently by two authors. The following information was collected: the first author's name, date of publication, country, study design, major CHD end point, selection of the controls, genotyping methods, allele frequencies of To determine the methodological quality of each study, we used the Newcastle-Ottawa scale (NOS). The NOS ranges between zero (worst) up to nine stars (best). Two authors of this article independently assessed the qualities of included studies. Disagreement was resolved by discussion. Apo E polymorphisms were estimated for each study with the software Stata12.0. The risk of the variant genotypes E2/2, E2/3, E2/4, E4/3, and E4/4 was estimated compared with the genotype E3/3 homozygotes. In addition to comparisons for total subjects, studies were categorized into different subgroup analyses according to the ethnicity. We estimated the between-study heterogeneity across the eligible comparisons using the Cochrane Q-test and the heterogeneity was considered significant for P < 0.1. Fixed effect or random effect was used to calculate pooled effect estimates. Random effects incorporate an estimate of the between-study variance and tend to provide wider confidence intervals, when the results of the constituent studies differ among themselves. In the absence of between-study heterogeneity, the two methods provide identical results.The risks  of CHD associated with theχ2 test to compare the observed genotypes frequencies with the expected ones among control subjects. For sensitivity analyses, we examined whether the excluding studies with substantial deviation from HWE among controls affected our pooled estimates of ORs.Publication bias was evaluated by funnel plot and Begg's and Egger's tests. The Hardy-Weinberg equilibrium (HWE) was tested by a goodness-of-fit K  is the number of included studies and Z is the Z value of the independent study. The result is obtained from the software SAS 9.2.Finally, we use the following formula to estimate the fail-safe number:714 studies were searched, among which 30 studies were included in the final meta-analysis –47. The P < 0.05). According to the quality criteria, there were 21 studies with high quality (NOS score > 6).Given in  Apo E alleles and genotypes in the cases and controls.Tables  ApoEε2 alleles and ApoEε4 alleles compared with the ε3 alleles in individual studies.Figures  ApoEE2/2, E2/3, and E2/4 genotypes did not have significant risk of CHD compared with individuals with the E3/3 genotype , However, compared with E3/3 genotype, the variant genotypes ApoEE4/4 and E3/4 were associated with increased risk of CHD in 30 studies in the genetic models .Overall, in the dominant model, individuals carrying the Apo E gene polymorphism and CHD risk in the genetic model of E3/4 and E4/4 compared with the genotype E3/3 in eight Mongolian studies , while the variant genotypes ApoEE2/2, E2/3, and E2/4 were not associated with the increased risk of CHD . In 22 Caucasians studies, significant associations were found in three genetic models ; carriers with ApoEE2/2 and E2/4 were not associated with the significant risk of CHD .In the subgroups, the results showed evidence of significant association between ApoEε2 allele had no significantly decreased risk of CHD compared with individuals with the ε3 allele in the random-effect model . Stratified analysis on the descent also showed no evidence on the ε2 allele variant and CHD risk in Mongolians , but there had a decrease risk in Caucasians . In addition, compared with the ApoEε3 allele, carriers with the ε4 allele had a 46% increased risk of CHD in the random-effect model , and the effects were more evident in the Mongolians  and mild in the Caucasians .Besides, carriers with Apo E polymorphisms and risk of CHD excluding studies deviating from HWE among controls. The pooled ORs estimates were similar with that of excluded studies, so the results were not shown.We conducted a sensitivity analysis on the ApoEε2 versus ε3 allele, the shape of the funnel plot seemed symmetrical (P = 0.211), and the fail-safe number also showed no publication bias in this meta-analysis (Nfs(0.05) = 381.51, Nfs(0.01) = 173.87).For themetrical , and the ApoEε4 compared with ε3 allele, the shape of the funnel plots seemed asymmetrical (P = 0.003). By using the trim and fill method, we showed that OR and 95% CI did not change. Besides, the fail-safe number also suggested that the result of the meta-analysis was stability (Nfs(0.05) = 1218.31, Nfs(0.01) = 588.44).For themetrical , and EggApo E) is one of the most major apolipoproteins in the central nervous system, with functions of neurons repair. Apo E genetic variants showed significant associations with the risks of nervous system degenerative diseases, including Alzheimer's disease, vascular dementia, and cerebrovascular disease [ Apo E gene as a receptor-binding ligand mediating the clearance of chylomicron and remnants of very-low-density lipoprotein cholesterol from plasma also plays a major role in the metabolisms of cholesterol and triglyceride. Functional variants of genes encoding lipoproteins are responsible in part for between-individual variation in the plasma levels of lipoproteins.Apolipoprotein E  compared with those carrying ε3 allele, while those with ε4 allele had a significant increased risk for CHD  [A lot of epidemiologic studies have investigated the relation between disease . The las ApoEE3/4 and E4/4 genotypes compared with carriers of the E3/3 genotype, but no significant evidence was found between the variant genotypes of ApoEE2/3, E2/4, and E2/2 and CHD risk. In the stratified analyses by descent, for the ε4 allele genetic variant, the effect was more evident in the Mongolians group and mild in Caucasians group, which showed that the ApoEε4 allele genetic variant modulated the increased risk of CHD with the differences of genetic background. Moreover, there was a decreased risk in Caucasians between the ε2 allele variant and CHD risk but no evidence in Mongolians; further studies should be conducted to verify it.In this 30 studies' meta-analysis including 11,804 CHD patients and 17,713 controls, we identified a significant increased risk for CHD among carriers of the ApoEε4 compared with ε3 allele, the Egger test showed existence of publication bias, but from the results of the trim and fill method and the fail-safe number, the publication bias and the possibility of false positive were relatively small. Second, the control group of some studies was not in conformity with Hardy-Weinberg equilibrium. But, in sensitivity analysis, the pooled estimates were similar after we excluded studies deviating from Hardy-Weinberg equilibrium among controls; therefore these studies were included in the final analysis. Gene-environment interactions may have contributed to the CHD. Apo E gene is a candidate gene and a common one to study gene-environment interactions. However, because of lack of the original data of the meta-analysis, further evaluation of potential gene-gene and gene-environment interactions was limited.Our study has several limitations. First, as with all meta-analyses, although we did Egger's test and calculated the fail-safe number to evaluate the publication bias, it might have occurred because our analyses were all based on published studies. For the ApoEε4 allele polymorphism may contribute to the individual susceptibility of CHD. Further rigorous design, large sample of case-control, or prospective study are required to continue in-depth evaluation on gene-gene and gene-environment interactions on Apo E polymorphisms and CHD risk.In conclusion, it was showed in this meta-analysis that


Opisthorchis viverrini, in the lower Mekong region. Numerous public health interventions have focused on reducing exposure to O. viverrini, but incidence of CCA in the region remains high. While this may indicate the inefficacy of public health interventions due to complex social and cultural factors, it may further indicate other risk factors or interactions with the parasite are important in pathogenesis of CCA. This systematic review aims to provide a comprehensive analysis of described risk factors for CCA in addition to O. viverrini to guide future integrative interventions.Cholangiocarcinoma (CCA) is a fatal bile duct cancer associated with infection by the liver fluke, O. viverrini. Seventeen risk factors were reported by two or more studies and were assessed with random effects models during meta-analysis. This meta-analysis indicates that the combination of alcohol and smoking  is most significantly associated with increased risk for CCA and is an even greater risk factor than O. viverrini exposure. This analysis also suggests that family history of cancer, consumption of raw cyprinoid fish, consumption of high nitrate foods, and praziquantel treatment are associated with significantly increased risk. These risk factors may have complex relationships with the host, parasite, or pathogenesis of CCA, and many of these risk factors were found to interact with each other in one or more studies.We searched five international and seven Thai research databases to identify studies relevant to risk factors for CCA in the lower Mekong region. Selected studies were assessed for risk of bias and quality in terms of study design, population, CCA diagnostic methods, and statistical methods. The final 18 included studies reported numerous risk factors which were grouped into behaviors, socioeconomics, diet, genetics, gender, immune response, other infections, and treatment for O. viverrini infection should be addressed in future public health interventions to reduce CCA in affected regions. In particular, smoking and alcohol use, dietary patterns, and socioeconomic factors should be considered when developing intervention programs to reduce CCA.Our findings suggest that a complex variety of risk factors in addition to The online version of this article (10.1186/s40249-018-0434-3) contains supplementary material, which is available to authorized users. Please see Additional file O. viverrini [O. viverrini is a foodborne helminth spread by ingestion of infected raw or undercooked cyprinoid fish in traditional local dishes [O. viverrini infection prevalence is up to 70% in some areas, with estimates of up to 10 million human infections in the lower Mekong region [O. viverrini has been classified as a group 1 carcinogen in humans by the International Agency for Research on Cancer (IARC) since 1994 [Cholangiocarcinoma (CCA) is a malignant tumor of the biliary tract occurring in high incidence in the lower Mekong region, including Thailand, Laos, Cambodia, and Vietnam. Annual CCA incidence in these areas ranges from 93.8 to 317.6 per 100 000 people, with most cases being fatal within 1 to 2 years of diagnosis . CCA in iverrini , 6 whichl dishes . O. viveO. viverrini may contribute to CCA development through induction of host immune response and inflammation in the bile ducts over the course of decades [O. viverrini infection in humans by incentivizing behavioral modification to reduce consumption of traditional raw fish dishes and reduce defecation in rice fields [O. viverrini infection in the lower Mekong region still remains high [O. viverrini infection and CCA incidence in the Mekong region [Chronic infection with  decades , 8, 11. e fields , 13. Howins high . Challenins high . Recent g region , 13, 16.O. viverrini infection has long been believed to be the primary risk factor, but numerous studies have also focused attention on other risk factors for CCA independently or in conjunction with O. viverrini [O. viverrini infection between women and men does not match this difference [O. viverrini infection prevalence and CCA incidence in Thailand is not consistent for all regions and may indicate importance of other risk factors [iverrini , 18. Forfference , 18, pro factors . An exam factors . Past reFollowing Preferred Reporting Items for Systematic Reviews and Meta-Analyses (PRISMA) guidelines , five inO. viverrini endemic lower Mekong region. All references matching these basic criteria were evaluated in full text for inclusion in the final systematic review, based on fulfillment of all of the following a priori inclusion criteria: the study design includes 1) human patients in the O. viverrini endemic lower Mekong region, 2) CCA diagnosis, 3) a comparison group without CCA, and 4) examination of risk factors for CCA in addition to O. viverrini. Included references were evaluated to determine weaknesses in study design in terms of limitations, risk of bias, choice of study population, definition of CCA cases, matching of controls, sample size, and statistical methods to assess the overall quality of the final set of references.The titles and abstracts of all references were screened by at least two reviewers for relevance to identify studies that reported primary research involving risk of CCA in humans in the CI) were recorded for risk factors reported in two or more studies for meta-analysis. Characteristics of each study including selection of the study population, diagnostics, study design, consideration of confounding, sample size, and statistical methods were extracted to examine sources of bias and heterogeneity. If more than one control group was included in a study, data from the healthy control group was used for consistency across studies. During data extraction, if overlapping datasets from the same patients were encountered in multiple included studies, data from the most complete report was included to avoid duplication of patient data in meta-analysis.At least two reviewers evaluated the full text of each included reference and extracted data for factors that were reported to increase, decrease, or have no significant effect on risk of CCA. Case and control exposure data or odds ratio and 95% confidence interval  was used for statistical analyses [Random effects models of log odds ratios were used to estimate summary measures for risk factors reported comparably by two or more studies. Random effects models were chosen to account for heterogeneity across included studies since heterogeneity was expected due to factors such as differences in patient source, diagnosis of CCA, measurement of exposures, and sample size. The Ianalyses .O. viverrini antibody response), gender, other infections, or treatment for O. viverrini infection . Increased alcohol consumption significantly increased risk of developing CCA, reported in five studies  [OR = 1.33, 95% CI: 1.00–1.78, P = 0.049). Meta-analysis of two studies reporting combined risk from alcohol consumption and smoking [OR = 11.1, 95% CI: 5.63–21.92, P <  0.0001). Meta-analysis of data from three studies [OR = 1.93, 95% CI: 1.2–3.1, P = 0.0065). Education was the only socioeconomic factor reported by two or more studies, and four studies [OR = 0.46, 95% CI: 0.22–0.97, P = 0.04). A family history of cancer, reported by three studies [OR = 3.0, 95% CI: 1.79–5.04, P <  0.0001). Raw fish dishes, which are vectors of O. viverrini, significantly increased risk of developing CCA [OR = 3.26, 95% CI: 1.58–6.71, P = 0.0014). High nitrate foods, inclusive of fermented or salted fish and meat and betel nut, also significantly increased risk for developing CCA . Though fresh fruit and vegetable consumption significantly reduced risk for developing CCA in some studies, the pooled effects in meta-analysis were not significant. Betel nut, fermented fish, rice, and sticky rice were also examined during meta-analysis with no significant association with CCA development  , 25, 26. smoking , 26 indi studies , 25 foun studies –24 indicAssessment of individual studies indicated that the majority of studies evaluated in this systematic review used a case control study design, so systematic recall bias must be considered since CCA is a cancer that develops over years of exposures. Patient reports may also be influenced by desirability bias based on what the patient perceives to be a desirable answer to report to physicians or researchers. Hospital or biological repository based sampling used in several studies also introduces bias, especially when control patients are selected from ill patients entering the hospital instead of healthy people in the community. Criteria for diagnosing CCA were variable between studies, with several allowing presumptive diagnosis via ultrasound examination instead of the gold standard by histology. Two studies used Japanese patients in the control group , 35 and O. viverrini as the primary, independent risk factor.To our knowledge, this is the first systematic review and meta-analysis to provide a comprehensive assessment of other risk factors for CCA described in the published literature and provides support for integrated interventions that consider the complexity of factor interrelations in the pathogenesis of CCA. Some of the described risk factors are primary and likely to be directly involved in the mechanism of pathogenesis of CCA, while others are secondary and may interact with or modify primary factors or represent a group of risk factors. Overall limitations of this systematic review to consider in light of these findings are the systematic recall bias and differences in measurements of exposures likely to be present across studies, as well as publication bias toward studies focused on O. viverrini exposure as measured by antibody titers


Low molecular mass protein (LMP) gene performs a critical role in the foreign antigen processing machine via the major histocompatibility complex-I (MHC-I) complex CD8+ cytotoxic T lymphocytes (CTL) pathway. Recent studies have reported the association of LMP2-60 G>A (rs17587) and LMP7-145 C>A (rs2071543) polymorphisms with various types of cancers, but the outcomes remained inconsistent. To obtain a reliable conclusion, we summarized available data and conducted a meta-analysis involving a total of 19 published studies. Evidences were obtained from the PubMed, Google Scholar, Web of Science and Chinese National Knowledge Infrastructure (CNKI) databases. The results demonstrated that the rs17587 and rs2071543 polymorphisms were associated with an increased cancer risk in the recessive and homozygote models. Stratified analyses by ethnicity indicated a significant association only in Asian population. Furthermore, rs17587 showed a greater susceptibility to gynecological cancers, while rs2071543 increased the risk of gastrointestinal and gynecological cancers. Our results indicate that the LMP2 rs17587 and LMP7 rs2071543 polymorphisms may act as risk factors for cancer, especially for Asian populations. Additional larger-scale multicenter studies should be performed to validate our results. Currently, cancer is a major cause of human death and a public health problem that seriously threatens human health worldwide . Large eIn recent years, many researches have confirmed that the elimination of cancer cell is promoted by the classical MHC-I-restricted T lymphocytes pathway , 3. In tPrevious researches indicated that the polymorphisms of LMP2- 60(G>A) and LMP7-145(C>A) would give rise to the functional alteration, and then impaired the capacity of antigen processing machine. As a consequence, LMP2/LMP7 gene variants were associated with development, occurrence and prognosis of many types of cancers, such as colorectal cancer, cervical cancer, gastric carcinoma and so on . HoweverThe flow chart of study selection process is shown in Figure P = 0.931, I2 = 0.0%; AA vs. GA/GG: OR = 1.31; (95% CI: 1.03-1.65), P = 0.787, I2 = 0.0%]. Subgroup analysis by ethnicity showed a significant association only in the Asian populations  , P = 0.446, I2 = 0.0%] , P = 0.572, I2 = 0.0%] , P = 0.506, I2 = 0.0%] , P = 0.679, I2 = 0.0%], but not in the Caucasian populations , P = 0.679, I2 = 0.0%] and gynecological cancers  , ] Figure , Table 2s Figure , Table 2s Figure , Table 2] Figure , Table 2P > 0.1) . The analysis outcomes demonstrated that no single study greatly influenced the overall cancer risk estimations with respect to the LMP polymorphisms. Figure , suggestWe also performed Begg’s and Egger’s test to assess the possible publication bias. The funnel plots of Begg’s test were symmetrical inverted Figure , which sCancer cells can escape immune recognition via insufficient expression of peptides presented by the MHC, because HLA-mediated presentation of immunogenic cancer peptides class I is a prerequisite of a successful antitumor immune response . Recent As a significant cause of population diversity, SNP is the most common type of human genetic variation correlated with cancer susceptibility . RecentlThe current study is the very first meta-analysis of the association between LMP2/LMP7 polymorphisms and cancer risk. Analysis among all subjects suggested a significant increase in the risk of cancer associated with LMP2/LMP7. In our meta-analysis, the genetic heterogeneity between the selected studies was evaluated, and no significant heterogeneity was observed. All studies included were found homogeneous without any study disproportionately driving the combined estimates. In subgroup analysis of ethnicity, our results showed that both rs17587 and rs2071543 were correlated with cancer risk in Asian populations, but not in Caucasian populations. The differences between Asians and other races may be partly attribute to different genetic backgrounds, lifestyles or environments. Besides, when the stratification analyses were conducted by cancer types, we identified a significantly increased susceptibility to gynecological cancer for rs17587 polymorphism, to gastrointestinal and gynecological cancer for rs2071543.Meta-analysis is a powerful method for analyzing cumulative data of studies where the single sample size is small and the statistical power is low . HoweverGenerally speaking, our meta-analysis manifests that LMP2/LMP7 polymorphisms  are risk factors for cancer, and presence of the two polymorphisms in Asian population will increase their susceptibility to cancer. However, larger-scale multicenter studies are needed to further clarify the possible roles of the polymorphisms in the development of cancers.The PubMed, Web of Science, Google Scholar and the China National Knowledge Infrastructure (CNKI) databases were searched for relevant studies up to April 14, 2017 in both English and Chinese through with the following terms and their combinations: “LMP2”, “LMP7”, “low molecular mass protein”, “polymorphism” or “variant”, “rs17587”, “rs2071543” and “cancer” or “tumor”. The reference lists of the retrieved studies were also screened to prevent the loss of any important data. All studies involved had to satisfy the following criteria: (a) case-control design was utilized; (b) studies focused on the association of LMP2/LMP7 polymorphisms with the risk of cancer; (c) published data must have been sufficient to allow OR estimation an odds ratio (OR) with a 95% CI. (d) for publications reporting the same data or overlapping data, only the largest or latest one was selected.Initially, all the following information was extracted independently by three investigators  and recorded in a standardized form. We extracted the following items from each article: including: first author’s name, year of publication, ethnicity of each study population, genotyping method, source of controls, cancer types, number of cancer cases and controls, allele frequencies, genotype distributions of LMP2/LMP7 polymorphisms, and Hardy-Weinberg equilibrium (HWE) results Table . DisagreThe studies quality was evaluated utilizing Newcastle-Ottawa Scale (NOS) for nonrandomized studies, including case-control and cohort studies . NOS awaAll analyses were performed using the Stata software, version 12.0 . All P values were 2-sided and a P value <0.5 was regarded as statistically significant.I2 test, with a higher I2 value indicating a higher level of heterogeneity . Meanwhile, if heterogeneity P value was higher than 0.10, the fixed effects model (the Mantel-Haenszel method) was used [To evaluate the cancer risk associated with LMP2/LMP7 polymorphisms, the pooled odds ratios (ORs) and 95% CIs were calculated. Heterogeneity between studies was evaluated using the was used . Otherwiwas used . Then, wA sensitivity analysis was performed to assess the stability of final results by calculating the outcomes again by omitting one single study at a time. Begg’s funnel plots and Egger’s linear regression were adopted to evaluate the publication bias . HWE was