To identify the association between gastrointestinal carcinomas (GIC) risk and UDP-glucuronosyltransferases (UGTs) 1A7 polymorphisms through a systematic review and network meta-analysis.Seventeen studies were eligible, which included 7738 patients and 18 analyses. First, it was found that compared with non-cancer participants, UGT1A7*1 were significantly decreased in cancer patient groups, especially in hepatocellular carcinoma, colorectal carcinoma, and Asian population groups; UGT1A7*2 was significantly increased in hepatocellular carcinoma and Asian population groups; and UGT1A7*3 was significantly increased in hepatocellular carcinoma, colorectal carcinoma, Caucasian, and Asian population groups. Second, the UGT1A7 polymorphism alleles contrast model and the categorized UGT 1A7 genotypes were compared, and the outcomes revealed that the ratio of UGT1A7*3 vs *2 increased, which may indicate an increased risk for cancer, especially for the pancreatic carcinoma and Caucasian groups. The ratio of Intermediate vs Low increased as well, which may also indicate an increased risk for GIC.PubMed, Embase, and the Cochrane library were searched for publications up until May 2017. First, the UGT 1A7 gene polymorphisms genotype in GIC patients were compared with a non-cancer control group, and second, the UGT1A7 polymorphism alleles contrast model and UGT 1A7 genotypes categorized according to enzymatic activity were examined.There is a cancer risk associated with increased UGT1A7 *2 for the hepatocellular carcinoma and Asian groups and with increased UGT1A7 *3 for the hepatocellular carcinoma, colorectal carcinoma, Caucasian, and Asian groups. Moreover, in Caucasian patients with GIC, the ratio of UGT1A7 *3 vs *2 was increased. Gastrointestinal carcinomas (GIC) present an increasing global public health threat, including several types of cancer, such as hepatocellular, colorectal, gastric, esophageal, and pancreatic cancer. GIC account for approximately 30% of all cancers worldwide, and most of them are characterized by a remarkable predominant incidence in males –2. HepatFor primary GIC with a hidden onset, rapid development, and a high degree of malignancy, an early clinical diagnosis is difficult. Many patients with symptomatic treatment are diagnosed in the late stages of cancer, and most have local metastasis and distant metastasis. GIC cancer is not sensitive to radiotherapy or chemotherapy, and a postoperative relapse is extremely common, which are the major reasons for a high mortality rate –7. Altho129R131W208), *2 (K129K131W208), *3 (K129K131R208), and *4 (N129R131R208). A haplotype analysis revealed that the polymorphisms at position 129 and 131 are in complete disequilibrium linkage, whereas the polymorphism at position 208 occurs independently. Based on these data, two polymorphisms  were detected for the UGT1A7 genotypic. UGT1A7 *3 and UGT1A7 *4 produce a lower corresponding catalytic activity for several substrates, including the B[a]P metabolites, when compared with the wild-type UGT1A7 *1 encoding enzyme (see Human UDP-glucuronosyltransferases (UGTs) are an enzyme superfamily, which could catalyze the glucuronidation of a diverse range of compounds, including endogenous metabolites , therapeutic drugs, and various classes of chemical carcinogens  –11. Polyzyme see  13]..129R131WNumerous valuable epidemiologic studies have suggested that UGT1A7 affects individual susceptibility to various carcinomas, such as HCC , CRC 1515, and Pn = 7738) and four UGT1A7 alleles were included . The P value was calculated using a meta-regression, and the results revealed that cancer type  might influence heterogeneity, with P = 0.006. A similar effect size was found in the subgroups HCC, CRC, and Asian population. Little evidence of bias could be found by the Begg’s test and Egger’s test, with a low to moderate quality of evidence according to the GRADE assessment.Figure P = 0.000, I2 = 68.7%), while the different types of cancer could still be heterogeneous sources by meta-regression (P = 0.012). Little evidence of bias was found by the Begg’s test and Egger’s test, and there was low-quality evidence according to the GRADE assessment.In addition, a significant increase was found only in the HCC group and Asian population group when UGT1A7*2 was compared in the cancer patient group and the control group. In general, an increasing trend was found in the GIC group  accompanied by substantial heterogeneity . Similar results were found in the subgroups of GCC, CRC, proximal digestive tract cancer, Caucasians, and Asians. The origin of heterogeneity could be different cancer types (P = 0.039) or ethnicities (P = 0.023) based on the meta-regression. Little evidence of bias was found by the Begg’s test and Egger’s test, and the evidence was low-quality according to the GRADE assessment.Moreover, the increase in UGT1A7*3 may be associated with a GIC risk  with substantial heterogeneity . The meta-regression did not reveal the source of heterogeneity.Finally, UGT1A7*4, which may be associated with an increased risk of cancer , was accompanied by low heterogeneity (P < 0.05). Thus, further comparisons using pairwise meta-analyses and network meta-analyses between the UGT1A7 polymorphism alleles contrast model and the categorized UGT 1A7 genotypes are necessary.In general, the pairwise meta-analysis suggested that the UGT1A7*3 allele was a risk factor for GIC. In addition, this phenomenon was found to be more prominent in some cancer type subgroups, such as HCC and CRC, and some ethnic subgroups, such as Caucasian and Asian populations; however, this conclusion was not sufficiently systematic, and the meta-regression results also showed that the differences between the two groups varied  and Intermediate vs Low  was found. In addition, for the Asian population, a significant cancer risk from Intermediate vs Low  was identified.The network meta-analysis represents the most comprehensive synthesis of data for currently available data for UGT1A7 polymorphisms and gastrointestinal cancer risk. Direct and indirect trials comparing UDT1A7 polymorphic alleles and categorized UGT1A7 genotypes that were reported for 7738 participants were reviewed. First, it was found that compared with non-cancer participants, UGT*1 was significantly decreased in cancer patient groups, especially in the HCC, CRC, and Asian population groups; UGT*2 was significantly increased in the HCC and Asian population groups; UGT*3 was significantly increased in the HCC, CRC, Caucasian, and Asian population groups. Secondly, the UGT1A7 polymorphism alleles contrast model and the categorized UGT 1A7 genotypes were compared, and the outcomes revealed that the increased ratio of UGT*3 vs *2 may indicate an increased risk of cancer, especially in the PC and Caucasian groups. The ratio of Intermediate vs Low increased as well, which may also indicate an increased risk of cancer.In recent years, the relation between gene polymorphisms, such as EGF, NAT2, DPYD, and UGTlAl, and the risk of GIC have attracted considerable attention. The UGT enzyme belongs to the super-gene family, which are the most important II metabolic enzymes in the human body. The catalytic glucoside reaction can catalyze alcohol, phenol, hydroxylamine, carboxylic acid, amides, mercaptan, and other types of chemical toxicities of the glucuronic acid binding reaction inactivation, and the detoxification mechanism plays an important role. UGTlA7 is a subtype of UGT, which changes metabolic enzyme activity due to mutations in genotypes and affects its ability to carry out detoxification, thereby altering the susceptibility of the body to carcinogenic factors .This study extends the findings from primary case-controlled trials and previous meta-analyses by systematically synthesizing the efficacy data –35. The This review followed the guidelines for conducting rigorous systematic reviews and network meta-analyses –41. To iThese results were proposed by the meta-analysis for the first time, and there is no published evidence to support this finding; however, it is logical to assume that a mutant type *3 with decreasing enzyme activity (compared with type*1 wild type) vs a mutant type *2 with a similar enzyme activity could increase cancer risks. Each time a normal cell divides, several errors or mutations occur. Most of these mutations do not cause harm because they occur in junk DNA, in genes unrelated to cancer, or in unimportant areas. Therefore, not all mutant-type genes are harmful. Though this is usually the case, which is favorable, this can cause harm when cancer-driven genes are involved . It is sThe network meta-analysis had some limitations that merit further discussion. First, in the GRADE framework, several comparisons were determined to be moderate or low-quality, which largely restricts the interpretation of the results. In addition, the network analysis contained some inconsistencies, which were mainly determined by the loop. Moreover, only four alleles were measured, and some alleles (such as UGT1A7 *5–*11) were not analyzed due to the small sample size. Furthermore, positive results are likely to be published, while negative results are not likely to be shared. An additional limitation of the pairwise outcomes was the extensive heterogeneity Tables , 3, whicThe findings of this comprehensive network meta-analysis provide some evidence that there are cancer risks associated with increased UGT1A7 *2 for the HCC and Asian groups and with increased UGT1A7 *3 for the HCC, CRC, Caucasian, and Asian groups. Moreover, in Caucasian patients with GIC, the ratio of UGT1A7 *3 vs *2 was increased. Genotype studies revealed that increased UGT1A7* 3 vs *2 ratios were the most likely to result in GIC, especially in the Caucasian population. In a clinical setting, the genetic monitoring of UGT1A7 can be used as a predictor of cancer.This systematic review was performed with an a priori established protocol (PROSPERO CRD42017064826) , and theThe inclusion criteria consisted of: case-controlled trials of subjects with and without gastrointestinal carcinomas ; participants of any age, gender, tumor stage, and histological grade; either Caucasian or Asian; and either using PCR as a genotypic method or DNA sequencing for detection. The exclusion criteria were: background records of comparing UGT 1A7 gene polymorphism genotypes in gastrointestinal cancers and an inappropriate control in the studies. Moreover, studies that used inappropriate tumor types (non-GIC) and that had no specific data, no human subjects, and no usable data were excluded.Three researchers  independently extracted data on studies, participants, and genotype-related characteristics to a standardized form, and discrepancies were resolved by consensus by referring to the original study in consultation with a fourth reviewer (LBA or ZQC). Data on UGT 1A7 gene polymorphism genotypes were extracted from original studies. Cancer type, trial size, and details of genetic polymorphisms, including UGT1A7 allele frequency and UGT1A7 categorized genotype, were also extracted.To reduce the risk of bias of individual studies, the Newcastle-Ottawa scale score  was usedSubgroup analyses and a meta-regression were performed according to cancer type  and ethnicity .The primary outcome was that a gastrointestinal cancer risk was associated with UGT1A7 polymorphism alleles , which could be stratified by cancer type and ethnicity. The secondary outcome was the UGT1A7 polymorphism alleles contrast model and UGT 1A7 genotypes categorized according to enzymatic activity , which could also be stratified by cancer type and ethnicity.P value and I2 statistic, with values over 50% indicating substantial heterogeneity [Studies that reported multiple cancer types and ethnicities were categorized as sub-studies (marked as a/b) to avoid double counting and mistreating data. First, a direct meta-analysis was performed using a random effects model because it is likely the most appropriate and conservative methodology to account for between-trial heterogeneity within each comparison –44. To eogeneity . The Begogeneity .P value and the I2 statistic.Second, a random-effects network meta-analysis was conducted using STATA v14.0. The results of the network meta-analysis were summarized using OR and their credible intervals (CrI) . A commoEach relative UGT1A7 polymorphism allele and UGT1A7 genotype resulted from the combination of the direct and the indirect evidence derived from the network meta-analysis, which was assumed to be coherent . InconsiIn addition, the quality of evidence for the primary outcomes was assessed based on the GRADE system using GRADEpro GDT –50. The No patients were involved in forming the research question or the outcome measures, nor were they involved in developing plans for the design or the implementation of the study. No patients were asked to advise in the interpretation or writing of the results. There is no intention to disseminate the results of the research to the study participants or to the relevant patient community.


P = 0.100). However, after excluding two studies for its heterogeneity and publication bias, no significant relationship was also detected between NAT2*4 allele and the increased risk of PCa, in fixed-effect model . Meanwhile, no significant increased risk of PCa was found in the subgroup analyses by ethnicity, source of controls and genotyping method. Moreover, TSA demonstrated that such association was confirmed in the present study. Therefore, this meta-analysis suggested that no significant association between NAT2 polymorphism and the risk of PCa was found.Previous studies have investigated the association between NAT2 polymorphism and the risk of prostate cancer (PCa). However, the findings from these studies remained inconsistent. Hence, we performed a meta-analysis to provide a more reliable conclusion about such associations. In the present meta-analysis, 13 independent case-control studies were included with a total of 14,469 PCa patients and 10,689 controls. All relevant studies published were searched in the databates PubMed, EMBASE, and Web of Science, till March 1st, 2017. We used the pooled odds ratios (ORs) with 95% confidence intervals (CIs) to evaluate the strength of the association between NAT2*4 allele and susceptibility to PCa. Subgroup analysis was carried out by ethnicity, source of controls and genotyping method. What's more, we also performed trial sequential analysis (TSA) to reduce the risk of type I error and evaluate whether the evidence of the results was firm. Firstly, our results indicated that NAT2*4 allele was not associated with PCa susceptibility (OR = 1.00, 95% CI= 0.95–1.05;  Prostate cancer (PCa) is considered as the second most leading cause of cancer-related deaths among men worldwide , 2. PCa N-acetyltransferase 2 (NAT2) gene is clustered on the short arm of chromosome 8 (8q22) and encodes a phase II xenobioticmetabolizing enzyme , 7. BesiTo date, a number of studies have been performed to elucidate the association between NAT2 genetic polymorphism and susceptibility to PCa. Meanwhile, some studies have demonstrated that people with low NAT2 activity have a higher risk of developing PCa, compared to those with high NAT2 activity . HoweverOverall, thirteen independent case-control studies were included with a total of 14,469 PCa patients and 10,689 controls in the present meta-analysis –34. TablP = 0.100). However, in order to find possible factors, heterogeneity analysis and publication bias were tested. After excluding two studies of Hamasaki et al. and Wang et al. in the existence of heterogeneity and publication bias, the results demonstrated that no significant relationship was also found between NAT2*4 allele and the risk of PCa, in fixed-effects model   Figure . MoreoveP = 0.131)  Figure .P = 0.451), which indicated that two studies by Hamasaki et al. and Wang et al. might have generated the origin of heterogeneity  An independent case-control design; (b) The relationship between NAT2 gene polymorphism and susceptibility to PCa was evaluated; (c) The abundant data to evaluate the pooled odds ratios (ORs) with 95% confidence intervals (CIs) was provided.Exclusion criteria from this meta-analysis were as follows: (a) Non-case-control studies; (b) Review articles; (c) No original available data of genotype frequency to evaluate this association; (d) Previous duplicated publications.Two investigators (Wang F and Qin ZQ) independently extracted the data from the identified studies. In addition, any disagreement was resolved by a discussion with a third reviewer until consensus was based on the main point of view. The following information were recorded in a standardized form: first author's last name, year of publication, country, ethnicity, source of controls , genotying method, number of cases and controls, and genotype frequency of NAT2 gene polymorphism between cases and controls, respectively.The quality of the studies was assessed using the validated Newcastle-Ottawa Scale (NOS) for non-randomized studies, including case-control and cohort studies. A study can be awarded a maximum of one star for each point within the selection and exposure categories, and a maximum of two stars can be given for comparability. We considered studies with scores of more than 7 as high-quality studies and only high-quality studies were included in our meta-analysis.P values < 0.05, the random-effects model would be conducted; Otherwise, the fixed effects model would be used to perform meta-analysis. What's more, subgroup analysis was further carried out to explore the potential sources of heterogeneity according to ethnicity, source of controls and genotyping methods. Besides, the influence of publication bias between the studies was estimated by Begg's funnel plots and Egger's linear regression test, and P < 0.05 was considered to be statistically significant. All above statistical data were conducted by Stata software .The pooled ORs with 95% CIs were calculated to assess the strength of association between NAT2*4 allele and PCa risk. Two models were used in the meta-analysis, including the fixed-effects model (a Mantel–Haenszel method) and the random-effects model (a DerSimonian–Laird method). If A cumulative meta-analyses with addition of new publishing trials might result in type I and type II errors, because it could increase the risk of random error with repeated significance testing and sparse data –17. TherThe results of this meta-analysis demonstrated that no evidence supporting the relationship between NAT2 polymorphism and PCa risk was detected. More importantly, further studies in other ethnic groups, such as Asians, Africans, are needed to give more comprehensive understanding of such association.


P = 0.292). We noticed that patients in Asia or Africa  and whose age less than 30 years  had substantially higher CRP levels. Our study provides evidence that higher CRP levels are associated with increased risk of SZ, especially for young adult patients less than 30 years. Further large-scale studies are strongly warranted to further confirm this association.C-reactive protein (CRP) has been indicated to be associated with the pathogenesis of schizophrenia (SZ) and other psychiatric disorders. The aim of this study is to investigate whether peripheral blood CRP levels are associated with the risk of SZ. We searched literature from databases of Pubmed, Embase and the Cochrane Library from inception to November 1, 2016 for studies that reported serum or plasma CRP levels in patients with SZ and non-SZ controls. At least two reviewers decided on eligibility and extracted data from included studies. Random effects meta-analyses were performed using standardized mean difference (SMD) as the effect estimate of the differences in CRP levels between subjects with SZ and healthy controls. We identified 18 studies representing 1963 patients with SZ and 3683 non-SZ controls. Compared with non-schizophrenics, blood CRP levels were moderately increased in people with SZ  irrespective of study region, sample size of included studies, patient mean age, age of SZ onset and patient body mass index. Publication bias was not detected through Egger's linear regression test ( Immune system dysfunctions and inflammatory processes have been implicated in the pathogenesis of schizophrenia (SZ) –5. A vasIt has also been indicated that the increase in serum or plasma CRP levels contributes the risk of a first episode of SZ . Inoshit2 and 11 studies with the range from 25 to 30 kg/m2. Most of the studies (17/18) applied high-sensitivity CRP assay except for one [The literature search yielded 927 citations. After excluding the duplicate, 39 appeared relevant and were retrieved for full-text review afer reading the titles or abstracts Figure . In summ for one . Twelve 2 = 92.5%, P < 0.001)  0.53, 95% CI 0.30 to 0.76), with significant heterogeneity among studies  and significantly in Asian/African countries . We found statistically significant difference for inter-study heterogeneity (P < 0.001).Subgroup analysis stratified by study region showed that serum and plasma CRP levels were increased moderately in studies conducted in European countries  and slightly in population-based samples . We found statistically significant difference for inter-study heterogeneity (P < 0.001).Meta-analysis stratified by clinical setting demonstated that serum and plasma CRP levels were increased moderately in studies performed in combination with inpatient and outpatient unit ; similar results were also obtained for studies with small sample size (< 100) .We found no statistically significant difference for inter-study heterogeneity (P = 0.075).Subgroup analysis stratified by sample size indicated that serum and plasma CRP levels were increased moderately in studies with large sample size (≥ 100) ; but moderately in studies with participants’age ranging from 30 to 50 years ; similar results were also obtained for studies with participants’age ≥ 50 years . We found statistically significant difference for inter-study heterogeneity (P < 0.001).Subgroup analysis stratified by age indicated that serum and plasma CRP levels were increased significantly in studies with participants’age <30 years ; but significantly in studies with participants’age ranging from 30 to 50 years ; similar results were also obtained for studies with participants’ age of SZ onset > 30 years . We found statistically significant difference for inter-study heterogeneity (P < 0.001).Subgroup analysis stratified by age of SZ onset showed that serum and plasma CRP levels were increased moderately in studies with participants’ age of SZ onset ≤ 30 years  and without adjusted variables for controls . We found no statistically significant difference for inter-study heterogeneity (P < 0.077).Subgroup analysis stratified by adjusted variables for controls showed that serum and plasma CRP levels were increased moderately in studies with . No significant funnel plot asymmetry was detected as was shown in Figure P = 0.544) and Egger’s regression test (P = 0.292). Duval and Tweedie’s trim and fill method indicated that no missing study was inputed and the adjusted SMD was the same as the primary one, confirming the robustness of the analysis.Sensitivity analysis showed that serum and plasma CRP levels were increased moderately in studies involved only high-sensitivity CRP assay  for the association between serum and plasma CRP levels and risk of SZ from the 18 studies included. The study quality of evidence obtained from NOS score was moderate and derived from case-control studies. However, still some other confounders remained due to factors that could not be accounted for such as physical activity, waist circumference, smoking habit, vitamin D intake, and subclinical infections. Most studies included in our analysis did not provide such data, which could explain the heterogeneity noted in the pooled analysis. Although subgroup analyses have been performed for some of the existing confounders, the investigated factors could only explain some of the heterogeneity. Moreover, many of the other analyses, such as dose-response relationship could not be conducted due to the lack of detailed information from original reports. Furthermore, the possible sources of heterogeneity was explored through subgroup analyses and sensitivity analyses. However, the results of these two methods did not allow us to attribute heterogeneity to any single study as the sole source of the high heterogeneity we obtained in most analyses. Secondly, this updated meta-analysis on serum and plasma CRP levels in patients with SZ compared with non-SZ controls provided us with summary estimates originating from one type of observational study (case-control study). Therefore, no definite causal conclusion can be drawn from this meta-analysis. High serum and plasma CRP level is only one of the inflammation-related factors known to be associated with the risk of SZ. Other factors such as smoking habit and subclinical infections could also be additional related factors. Thirdly, like any meta-analysis, this one is also dependent on the quality of the included studies, and our findings need to be further confirmed by studies specifically designed to demonstrate the unclear points we raised. Finally, the restriction of our meta-analysis to English language studies may have led to language bias with potentially relevant studies published in other languages being missed.In summary, our meta-analyses provide evidence that higher CRP levels are associated with increased risk of SZ, especially for young adult patients less than 30 years. Future studies should be focused on whether changes in CRP levels have a causal relationship with the development of SZ.The Preferred reporting items for systematic reviews and meta-analyses (PRISMA) statement was used when we performed this systematic literature review . We systStudies were considered eligible in the meta-analysis according to the following criteria: 1) adult patients diagnosed with SZ based on Diagnostic and Statistical Manual for Mental Disorder-Fourth Edition-Text Revised (DSM–IV–TR) were enrolled in the study; 2) case-control studies with the non-SZ comparisons measuring serum or plasma CRP levels. We excluded studies with duplicate data or studies that reported CRP levels using dichotomous data , or studies without a control group. In addition, we did not include unpublished literature due to the insufficient data obtained from these studies.Two authors (Z.W. and P.L.) independently searched, identified the related studies, extracted data and evaluated the study quality. When discrepancies occured, a third author (G.C.) made the definitive decision for study eligibility and data extraction. We extracted the following data including first author, publication year , research country, study subjects, clinical setting, sex, age and BMI of the study subjects, age of SZ onset, adjusted variables for controls, percent of the current smokers, CRP assay type, comorbidities and psychiatric drugs used. If necessary, we consulted the original data from the authors of the studies to collect missing information. The Newcastle–Ottawa Scales (NOS) was used to assess quality of the included studies , which a2 and Q test statistics, with an I2 more than 50% or a P value less than 0.1 from Cochran’s Q-test indicating significant heterogeneity [For continuous outcomes reported on different measurement methods for CRP levels, we applied standardised mean difference (SMD) estimates of the differences in CRP levels between patients with SZ and non-SZ controls as the effect size. The value of SMD being 0.2 was set to indicate a slight effect, meaning a small difference in CRP levels between patients with SZ and non-SZ controls, 0.5 a moderate effect, and 0.8 a significant effect . The hetogeneity . Summaryogeneity  and Eggeogeneity . We alsoogeneity . Some plogeneity . All met


Head injury is reported to be associated with increased risks of dementia and Alzheimer’s disease (AD) in many but not all the epidemiological studies. We conducted a systematic review and meta-analysis to estimate the relative effect of head injury on dementia and AD risks.Relevant cohort and case-control studies published between Jan 1, 1990, and Mar 31, 2015 were searched in PubMed, Web of Science, Scopus, and ScienceDirect. We used the random-effect model in this meta-analysis to take into account heterogeneity among studies.Data from 32 studies, representing 2,013,197 individuals, 13,866 dementia events and 8,166 AD events, were included in the analysis. Overall, the pooled relative risk (RR) estimates showed that head injury significantly increased the risks of any dementia  and AD , with no evidence of publication bias. However, when considering the status of unconsciousness, head injury with loss of consciousness did not show significant association with dementia  and AD . Additionally, this positive association did not reach statistical significance in female participants.The findings from this meta-analysis indicate that head injury is associated with increased risks of dementia and AD. The estimated annual global incidence of head injury requiring medical attention or resulting in hospitalization or death is over 10 million, and the risk of subsequent morbidity, mortality and disability is high.Neurodeget al. and  and 14] A study was eligible for inclusion if the following criteria were met: (1) examination of head injury or traumatic brain injury (TBI) as the variable of interest; (2) determination of incidence of AD or other types of dementias as the outcome of interest; and (3) reporting the RRs or odds ratios (ORs) of dementia or AD with their 95% confidence intervals (CIs). The studies about animal experiment, mechanistic research and review research were excluded.We extracted the characteristics of each included study, including author, study region, study design, sample size, mean age of the sample, exposure ascertainment, exposure variable, outcome (any dementia or AD), disease ascertainment, RRs or ORs with CIs, and factors adjusted for. The most adjusted estimate was included when a study reported more than one risk estimate, and crude risk estimate was included if a study did not adjust for other factors. For the purpose of sensitivity analysis, we also extracted information on minimally adjusted (crude or adjusted for sex and age) risk estimates from each study when available. The quality of each study was assessed by the Newcastle-Ottawa Scale recommended by Wells and colleagues.I-squared (I2) statistic and Q-statistic were used to explore the heterogeneity among studies. Large I2 (>50%) or P<0.10 for Q-statistic suggests substantial heterogeneity among studies. Subgroup analyses were performed according to the status of unconsciousness, including head injury regardless of status of unconsciousness, head injury with loss of consciousness (LOC), and head injury without LOC. We did several sensitivity analyses: mean age of the participants (≥65 year vs<65 years), sex, geographic location , study design (cohort or case-control), exposure type (TBI or other types of head injury), disease ascertainment methods, study quality score (full marks or not full marks), and year of publication (pre-2000 or 2000 onwards). We also performed sensitivity analyses by removing each individual study from the meta-analysis. We used funnel plots to examine the presence of publication bias . Egger’s regression test and Begg-Mazumdar test were used to further assess publication bias. All statistical analyses were done with Stata Version 12.0 software .A previous study indicated that OR is close to RR when the outcome is relatively uncommon (less than 20%). Thus, iThe systematic search identified 2887 articles, which were assessed by title and abstract. Of these, 91 articles were qualified for selection . After fI2 statistic for heterogeneity between studies was 91.2%, with p value for the Q test <0.001, suggesting substantial between-study heterogeneity. In the sub-group analyses, when regardless of status of unconsciousness, the pooled RR was 1.78 (95% CI 1.44–2.20); however, head injury with LOC did not show significant association with risk of dementia   . The I2 67–1.27) .I2 statistic for heterogeneity between studies was 70.1%, with p value for the Q test <0.001, suggesting substantial between-study heterogeneity. In the subgroup analyses, head injury regardless of status of unconsciousness , head injury with LOC  and head injury without LOC  were all associated with increased risk of AD  (The pooled RR estimates for risk of AD associated with head injury was 1.51(95% CI 1.26–1.80) . The I2 sk of AD . Exclusip<0.001) . Howeverp<0.001) .In the sensitivity analyses, the pooled RRs for any dementia and AD did not differ significantly by mean age of the participants, sex, geographic location, study design, exposure type of head injury, disease ascertainment methods, study quality score, or year of publication . HoweverP = 0.327 and P = 0.139, respectively) and Begg-Mazumdar test  further confirmed that there was no potential publication bias in both meta-analyses.Visual assessment of funnel plots showed that the studies were distributed fairly symmetrically about the combined effect size in both meta-analyses , which sIn this pooled analysis of 32 observational studies, with data for more than two million individuals and more than 13,000 dementia events and 8,000 AD events, head injury was a risk factor for any dementia and AD. Indeed, we observed an overall 63% increase in any dementia risk and 51% increase in AD, comparing individuals with head injury to those without head injury. These findings add to the accumulating evidence that head injury may lead to neurodegenerative diseases, although current evidence from the literature is mixed.et al assessed the relationship between TBI and subsequent neurological and psychiatric disease in a meta-analysis[et al’s study did not find significant association between TBI and dementia , which is inconsistent with our result. This might be due to the limited number of studies that included in their meta-analysis.The findings of this pooled analysis are partially consistent with the positive association observed between head injury and AD observed in a meta-analysis that combined case-control studies that were conducted up to the year 2001. The pre-analysis, using dOur meta-analysis included not only retrospective case-control studies but also prospective cohort studies. An important strength of prospective studies is that recruitment takes place and information about head injury is recorded before participants know whether they will develop dementia or AD. The robustness of prospective data is demonstrated by the stability of the findings in the sensitivity analyses. When the retrospective studies were assessed in isolation in the sensitivity analyses, their aggregate findings differed from those of the prospective studies, which may be caused by biases in some retrospective studies. Many retrospective study results could have been somewhat biased by selective participation of head injury patients, and in all retrospective studies information about head injury was recorded after dementia or AD diagnosis, so there might have been differential recall of head injury.An interesting finding in our subgroup analyses indicated that head injury with LOC had no significant association with dementia, and had no significant association with AD in the meta-analysis that excluded studies with results not adjusted for other factors. However, head injury without LOC showed consistently positive association with AD. This finding conflicted with the preconceived opinion that more severe injury induces more serious complications. One primary possible explanation was that most included studies did not distinguish head injury with and without LOC. Thus, there were very limited studies in the head injury with LOC or without LOC subgroup, making the results of subgroup low of statistical power. More studies are needed to further assess the relative risk of dementia/AD induced by head injury with LOC and without LOC separately. Another explanation was that head injuries without LOC would be susceptible to greater recall bias, and if that were so, one might observe a greater risk for AD among head injured persons without than those with LOC. Also, tIn the sensitivity analyses by sex, the positive association between head injury and dementia/AD did not reach statistical significance in female participants. The previous meta-analysis also indicated that the excess risk of head injury in those with AD is only found in males. The sexThe strengths of the present meta-analysis include that almost all the worldwide evidence from eligible epidemiological studies was included, as well as the acceptable methodologic quality of the studies on which the analysis is based. One limitation of our study is that all types of “head injury” were treated equally in the pooled analyses. Although we did sensitivity analyses by exposure type (TBI or other types of head injury), the results show that other types of head injury didn’t show significant association with increased risk of any dementia and TBI didn’t show significant association with increased risk of AD. This may be due to the limited number of studies included in the sensitivity analyses. More studies are needed to further clarify the relationship between a true TBI and the risk of dementia and AD. Other possible limitations of the study include the heterogeneity between the studies, including studies in which outcomes were recorded with different scales. Defining the status of head injury at baseline (cohort studies) or during the reference period (case-control studies) possibly caused the heterogeneity between the included studies, because the follow-up periods and the reference dates varied between the studies. Moreover, there were numerous differences in the analytic methods used to estimate the ORs and RRs among different studies, which may also contribute to the heterogeneity in the results. Therefore, our results must be interpreted with caution. We assumed that the true effect would vary between the studies because of potential additional heterogeneity in the populations, designs, and analyses of the various studies, in addition to the usual sampling variation in the estimates within studies. To account for the heterogeneity, we used the random effects model to combine the results of the original studies. The random-effect approach provides some allowance for heterogeneity in studies beyond sampling error. One can expect a very limited influence of heterogeneity by using the random effects model, although this does not necessarily rule out the effect of heterogeneity between the studies. Sensitivity analyses by some study-level factors were performed to further explore the sources of heterogeneity, but heterogeneity persisted after we performed relevant sensitivity analyses. Although the studies included in our meta-analysis were heterogeneous, the relation is largely consistent.Our meta-analysis is based on observational studies, and the possibility of selection bias, misclassification bias related to exposure, and failure to consider residual or unmeasured confounding cannot be ruled out. The assessment methods for head injury may also vary between the studies. The assessment in several included studies was based on self-reported questionnaire and medical records documenting the severity of injury were not accessed for all participants, and such data are subject to recall bias, especially for the patients with dementia and AD. Moreover, the present data do not address the important issues of whether a single head injury and repeated head injury can both increase the risk of dementia and AD, and whether a recent injury and an earlier injury on head or whether mild TBI and serious TBI have different impact on the risk of dementia and AD. Unfortunately, the data available in the current study do not allow for more refined categorisations of head injury. The time elapsed between head injury and dementia symptoms starting was reported rarely among the included studies, and this may be another limitation of this study. Besides, without autopsy confirmation, clinical diagnosis of AD is suspect due to all the “AD mimics”.Limitations aside, this meta-analysis remains the most comprehensive study to date that addressed the association between head injury and risk of dementia and AD. The diversity of location, ethnicity, age of participants, and head injury status reported in these studies also allows for increased generalisability of these results to other populations. However, studies included in our meta-analysis are mainly from Europe and The United States, thus, the results of our study must be interpreted with caution when generalized to populations from other regions. Although we recognize the methodological limitations of the studies included in this meta-analysis, our study does serve as a comprehensive review of this literature.On the basis of epidemiologic evidence, we found that head injury was associated with an increased risk of dementia and AD. For further studies, based on our findings, we suggest that the investigators should improve the standardization of various assessment methods of head injury, dementia and AD. Furthermore, this study adds to the existing evidence that head injury may lead to neurodegenerative diseases, and the use of genetic and biological makers as surrogate end points in the future studies should help to clarify the case and effect relationship that links head injury and dementia and AD.S1 Appendix(DOC)Click here for additional data file.S2 Appendix(TIF)Click here for additional data file.S1 FigBox sizes are in proportion to study weight. TBI = traumatic brain injury.(TIF)Click here for additional data file.S2 FigPooled relative risks for any dementia associated with head injury by omitting one study in turn.(TIF)Click here for additional data file.S3 FigPooled relative risks for Alzheimer’s disease associated with head injury by omitting one study in turn.(TIF)Click here for additional data file.S1 Table(DOCX)Click here for additional data file.S2 Table(DOCX)Click here for additional data file.S3 Table(DOCX)Click here for additional data file.


Larger and well-designed, preferably prospective studies are needed to further confirm our findings.The single nucleotide polymorphism (SNP) rs2910164 G>C within miR-146a has been reported that is associated with the increased risk of gastric cancer (GCa). However, the results are inconclusive, espicially among Asian populations, which probably due to small sample size in each single study. To validate this association and get a more precise estimation, we conducted a large GCa study including 1,125 cases and 1,196 controls in an eastern Chinese population. Our results showed that this SNP was not associated with GCa risk in either of the three genetic models . Stratified analysis by age, gender, smoking status, drinking status, or tumor location confirmed this non-significant association. In summary, these results suggest that the  In addimicroRNA-146a (miR-146a) is placed in the passenger strand, of which the C allele probably causes hairpins mispaired [miRNA, a small non-coding RNA molecule consisting of ~22 nucleotides, has crucial biological functions in post-transcriptional regulation of genes, as well as cell differentiation, proliferation, and apoptosis –12. All ispaired  and thusispaired . This miispaired –20. The ispaired , 22. Howispaired , 23–25, ispaired . HoweverTable Allele frequencies of the rs2910164 A>G SNP were listed by cases and controls in Table miR-146a has drawn an increasing attention for its potential connection to several types of cancers, including gastric cancer. Several studies have indicated the common SNP rs2910164 in miR-146a as a moderate risk allele for gastric cancer [Genetic susceptibility has been a research focus in cancer studies. Recently, c cancer , 28. HowWe acknowledge some limitations of the present study. First, although age, sex, smoking and drinking status, and tumor site were taken into consideration for subgroup analysis, other important risk factors, such as diet, microbial virulence, and HP infection, were missing in the study, which might also contribute to the etiology of GCa , 30. SecHP infection status or Lauren classification probably provide an updated result.In summary, these results suggest that the SNP rs2910164 of miR-146a may not be associated with the risk of GCa in this Chinese population. However, analysis of this SNP incorporating diet, This study included patients who were recruited from our ongoing molecular epidemiology study of GCa, and the cases and controls were described previously , 32, 33.According to a relevant protocol, we extracted genomic DNA from peripheral blood. The rs2910164 SNP was genotyped by the TaqMan assay with ABI7900HT real-time PCR system (Applied Biosystems) as reported previously . Partici2 test was used to assess differences in the distributions of demographic characteristics between cases and controls. The association between SNP and GCa risk was assessed by odds ratio (OR) and 95% confidence intervals (CIs) in heterozygous (CG vs. CC), homozygous (GG vs. CC), dominant (CG+GG vs. CC), recessive (GG vs. CG+CC), and additive (G vs. C) genetic models, respectively. OR values were calculated by both univariate and multivariate logistic regression models. Moreover, logistic regression tests for each genetic model were adjusted for age, sex, drinking and smoking status. Furthermore, the association between the miR-146a rs2910164 SNP and GCa risk was also stratified by age, sex, smoking or drinking status, and primary tumor site. All statistical process above was achieved by SAS Version 9.1 software .The χ


Phosphodiesterase type 5 (PDE5) inhibitors are recommended for patients with erectile dysfunction by American Urological Association and European Association Urology guidelines. However, recent researches have shown that PDE5 inhibitors may lead to increased melanoma risk. Thus, we aimed to explore whether PDE5 inhibitors are associated with increased melanoma risk based on published literatures.We conducted a systematic online search on PubMed, EMBASE, Cochrane Library, Chinese Biochemical Literature, China National Knowledge Infrastructure, and Chinese Science and Technology Periodical databases to identify the related studies. Odds ratios (ORs), risk ratios, and hazard ratios with 95% confidence intervals (CIs) were extracted and calculated to assess the strength of associations between PDE5 inhibitors and melanoma risk. We also extracted the basal cell carcinoma (BCC) to validate the association in this study.We included 5 studies containing 100,932 participants in our systematic review and meta-analysis. The calculated results suggested positive results of PDE5 inhibitors on melanoma risk . For localized and nonlocalized melanoma, the results were different  . It also showed that PDE5 inhibitors were associated with increased BCC risk .The association between PDE5 inhibitors and melanoma might not be causal due to potential bias  and limitations. With only mild complications reported, vision-threatening ocular complications, and hearing loss, PDE5 inhibitor become the first-line therapy for ED.–3 Sildenafil, tadalafil, vardenafil, and avanafil are 4 drugs approved by the US Food and Drug Administration specifically for ED treatment.Phosphodiesterase type 5 (PDE5) inhibitors are a kind of drugs which is widely used in clinic. By completely inhibit decomposability of cyclic guanosine monophosphate (cGMP), PDE5 inhibitors are expected to result in relaxation of smooth muscle and maintain penile erection. declared that sildenafil might induced invasion of melanoma in 2011. Since then, more and more concerns have been raised on this subject. An increased risk of skin melanoma following sildenafil use was also noted by Li et al in 2014 . Melanoma is the most aggressive type of skin cancers worldwide, especially in western countries with lighter skin colors. It only counts approximately 10% of all kinds of skin cancers, but it causes >80% of skin-related death. A possible explanation is the PDE5 is part of the RAS-RAF-MEK-ERK signaling pathway which has been involved in the development of melanoma. Also, the reduced PDE5 expression triggered by activation of BRAF gene activation would increase the invasiveness and metastatic potential of melanoma cells, which was noted in >50% melanomas and stimulates melanoma cell invasion and metastasis.,8,9 In addition, a recent published study showed sildenafil promotes melanoma growth by potentiating a cGMP-dependent pathway. It is thus plausible that direct pharmacological inhibition of a PDE5 activator may increase the risk of developing melanoma.While scientist focuses on kidney disease and PDE5 inhibitors, because of wide expression of PDE5 in tissues, Arozarena et al,12 Moreover, the patent of sildenafil and other PDE5 inhibitors have expired or are soon going to expire in various countries, which lead to the availability of less costly generic versions and the potential for considerably inflated demand in the near future. Therefore, it is important to figure out if there is an association exists between PD5 inhibitors and melanoma risk. However, since 2014, several studies have been published with conflicting results.,13–16 We attempted to investigate the association between PDE5 inhibitor use and melanoma risk with a meta-analysis. We also extracted the odds ratios (OR) and HR of PDE5 inhibitor use and basal cell carcinoma (BCC) to re-evaluate the validity of the association of PDE5 inhibitors and melanoma risk.Any increase in malignant melanoma risk that is caused by PDE5 inhibitors would have serious public health implications .22.1Two independent investigators, conducted a systematic search of Pubmed, EMBASE, Cochrane Library, Chinese Biochemical Literature (CBL), China National Knowledge Infrastructure (CNKI) and Chinese Science and technology Periodical (CSTP) databases to identify studies related to the association of PDE5 inhibitors and melanoma risk. Search terms were “Melanoma,” “Malignant melanoma,” “Phosphodiesterase type 5 inhibitor,” “PDE5,” “Sildenafil,” “Tadanafil,” “Vardenafil,” and “avanafil.” The references of included studies were also checked manually, in case of missing relevant studies. No language restrictions were applied in this meta-analysis, and the latest online search was in January 2017. The ethical approval of the present study is not necessary, because this is a meta-analysis which is based on published literatures. And no new human participants are involved in this study.The inclusion criteria were: studies related to the associations of PDE5 inhibitors and melanoma risk; randomized-controlled trials, cohort studies, or case–control studies; studies provided with OR with 95% CI and risk ratios (RR) with 95%CI or HR with 95% CI. Accordingly, case reports, abstracts, conference proceedings, reviews, letters, or repeated publications were excluded. We did not select the ethnicity of the study population. Studies identification, quality assessment, and data extraction were conducted by 2 individual reviewers. If any disagreement appears, a third reviewer was asked to help solving it. All the related articles were retrieved on the internet. If not available, we tried to contact the author directly for full articles.2.2 For nonrandomized controlled studies, we used the Newcastle–Ottawa scale (NOS) to assess the quality of the studies. Studies with a score >7 were considered to be highly qualified.The quality of each included study was evaluated using the Grades of Recommendation Assessment and Development and Evaluation (GRADE) approach.The basic information of included studies was extracted: name of the first author, year of study recruitment, age range, country, study design, study population data source, PDE5 inhibitor data source, number of patients, and outcome evaluation. Multivariable adjusted ORs, RRs, HRs, and their 95%CIs were also extracted.2.3 The pooled out unadjusted risk estimate were used in determining the strength of the association between PDE5 inhibitors and melanoma risk. We also extracted the ORs/RRs/HRs of PDE5 inhibitor use and BCC risk to validate of the association of PDE5 inhibitors and melanoma risk (the risk of BCC was not expected to be associated with PDE5 inhibitors). The heterogeneity test was performed using the χ2 test based on Q-test and I2-test. We used random-effects model throughout whole study since there is considerable heterogeneity in some meta-analyses , regardless of P value and I2. When 2-sided P <.05, we defined it as statistically different. Inverted funnel plot visual inspection was used to rate the publication bias among included studies. Besides that, we also performed a sensitivity analysis by excluding each study one by one and recalculate data of the remaining studies. All statistical analyses were performed by Revman software .The extracted data from included studies were unadjusted ORs/RRs/HRs with 95%CI except a few multivariable adjusted ORs/RRs/HRs. If the OR was not available nor given the original data, but presented with HR, we used the HR to represent RR. And RR can be transferred to OR using Willi's methods.33.1Our online systematic search identified 46 related studies using the search strategies described above. After screening the titles and abstracts of all studies, we excluded 36 irrelevant studies and 5 letters to the editors. In all, we got 5 studies qualified for our meta-analysis. A flow diagram of study selection was shown to present how we identify pertinent studies Fig. .The included studies were published between 2014 and 2016, containing 100,932 participants . All studies were conducted in western, developed countries . Three studies were cohort studies using HR, and the remaining 2 studies were case–control studies using OR. Only 1 study reported data from 2 data sources, while the other 4 studies reported results of 1 data source. The outcome evaluations of the studies are melanoma, BCC, and squamous cell carcinoma (SCC). All included studies were considered as high-quality studies via NOS evaluation. Sensitivity test did not show abnormal findings. All the summarized characteristics of the included studies are shown in Table 3.2I2 = 45%, P = .11), using a randomized model. . No statistically heterogeneity was noticed among 5 studies , while no statistically significant heterogeneity found  . The heterogeneity was 3.3I2 = 79%, P = .002) . However, a statistically noticeable heterogeneity was found (02) Fig. .4To the best of our knowledge, this is the first meta-analysis regarding the association between PDE5 inhibitors and melanoma risk. We included 5 studies containing 100,932 participants in this systematic review and meta-analysis. And, we found that PDE5 inhibitor use might associate with increased risk of melanoma, especially with localized melanoma and BCC, while decreased risk of nonlocalized melanoma. study, the exposure was assessed using questionnaires when starting the cohort in 2000, and update on exposure data was not obtained, and lacked important data on timing, duration, and dosing of PDEIs, tumor stage, and use of PDE inhibitors other than sildenafil. Loeb et al's study was limited to patients who had filled with a single prescription, and the proportion of younger men is lower compared to other studies which resulted in marginally lower power and wider confidence intervals for this subgroup. Lian et al using CPRD records prescriptions written by general practitioners and not those filled by patients, leading to some exposure misclassification. Matthews et al's study limitation is they did not have individual-level data on sun exposure, so no directly control for this in the primary analysis. Pottegard et al did not evaluate the BCC or SCC which as negative control of melanoma risk. Moreover, 4 out of 5 studies claimed the association might not be causal. All the studies reported above did not show the participation rate, thus, we do not know whether the participants can represent the general population which is quite important. Though there is no existing evidence on connection between PDE5 inhibitors and risk of BCC, the present study gave out a positive outcome. Moreover, we noticed that in Mathews et al's study, a similar increased risk was also noted in solar keratosis . They thought the possible explanation may lies in the sun exposure. Their study showed strong evidence that PDE5 inhibitor users are more likely to have solar keratosis before PDE5 inhibitors prescription, which proposed the possibility that those PDE5 inhibitor users might had experienced excess sun exposure and it is more likely to develop melanoma. Another evidence is that the increased risk was not recorded in the same study for colorectal cancer , which is not related to sun exposure. We all know that the common connection among melanoma, BCC, and solar keratosis is sun exposure. However, most included studies, in the current analysis, failed to evaluate sun exposure among participants or the evaluation was not reliable. It also should be addressed that in several studies, the education level and the income of 2 groups were statistically different. This might lead to different life-style between the 2 groups. Thus, the result may be affected by other values which were not be evaluated.The present studies indicated that there is an association between PDE5 inhibitors and melanoma risk. However, this association must be carefully interpreted, because it is not likely to be causal. Though 4 out of 5 studies used medical record as the definition of PDE5 inhibitor exposure which are highly reliable, all 5 included studies have limitations of one or another kind. In Li et al's and Matthews et al, though Matthews et al came up with positive result, the conclusions of both studies indicated that the association between PDE5 inhibitors and melanoma is not causal (due to sun exposure).When we compared the results between cohort studies and case–control studies, the overall results are still not consistent . The most recent cohort studies are conducted by Lian et al study, incorporated in the analyses as markers of healthcare-seeking behavior.The subgroup analysis suggested different results between localized and nonlocalized melanoma. We considered that may be compatible with a detection bias stemming from that PDE5 inhibitor users had more intensive contact with the healthcare system. The possibility of detection bias is also supported by the observed attenuation of associations after adjustment for education and ambulatory visits in Pottegard et al's HR: 0.84l; 95%CI: 0.59–1.20; Lian et al HR: 1.12, 95%CI: 0.87–1.44). SCC is also mechanically not associated with PDE5 inhibitors. Interestingly, 2 studies provided different outcomes of PDE5 inhibitors and melanoma risk. Lian et al's study showed no association, but Li et al's showed positive one. Since Li et al's study only included 14 melanoma patients used sildenafil which is less than Lian et al's study, this difference might be caused by potential bias and mislead the interpretation of the results.Excepted the results reported above, we also found some interesting results regarding other aspects. Nevertheless, these data cannot be combined due to lack of studies. There are 2 studies reported the PDE5 inhibitors use and risk of SCC. Both of the studies demonstrated no associations  and ever use . Regarding to lower dose, no statistically increased melanoma risk was found. However, in Lian et al's studies showed an increased risk among participants who had received ≥7 prescriptions or more than 25 tablets . On the other hand, in Lian et al's study, it seems PDE5 inhibitors will not increase or decrease the melanoma risk in fewer prescriptions (1–6) and lower dosage (below 25 tablets) . These results cannot be combined due to obviously heterogeneity. But considering its study design , the cohort studies provide evidences more powerful than case–control studies. Thus, it is hard to declare that there does exist a dose-dependent association between melanoma risk and PDE5 inhibitors. Also, the proportion of people diagnosed with melanoma differs from countries. According to the included literatures, the highest prevalence of melanoma lies in Loeb et al's study, which is 16.7% of the whole population, while the lowest is only 0.2% in Matthews study. So, this also could be the reason why these results are not consistent.Some studies assessed whether the dosage of PDE5 inhibitors would change the results. Pottegard et al's study. Usage of 200 to 499 tablets of tadalafil resulted in an OR is 2.05 (95%CI: 1.10–3.84), while the ORs were 1.44 (95%CI: 0.99–2.11) and 1.39 (95%CI: 0.58–3.32) for 200 to 499 and 500+ tablets of sildenafil use in the Danish Nationwide Health Registries (DNHR) database. In Kaiser Permanente Northern California (KPNC) population, the use of 500+ tablets of sildenafil was associated with an OR of 2.50 (95%CI: 0.91–6.99), and the use of 200 to 499 tablets of vardenafil suggested an OR of 1.38 (95%CI: 0.88–2.16). Loeb et al reported that the risk estimates were similar among sildenafil, vardenafil and tadalafil. These results indicated that there might be no statistical difference between different PDE5 inhibitors. Giving the preceding controversies, more high-quality studies are still required for further validation.Regarding PDE5 inhibitors of different types, namly sildenafil, tadalafil and vardenafil, different results were shown in Porttegard et al'sThere are several potential limitations in the present study. First, there are only 5 studies included in this meta-analysis. Though, the studies are all well designed, more studies are still needed especially when potential bias occurs. The case–control studies and cohort studies might bring potential biases when missing some matches of important values. adjusted for age, body mass index, smoking, physical activity, childhood reaction to sun, number of sunburns, mole count, hair color, family history of melanoma, and sun exposure; Pottegard et al adjusted for use of oral drugs like steroids, weak/moderate topical steroids, and diagnoses of nonmelanoma skin cancer, type 1 or type 2 diabetes, chronic obstructive pulmonary disease, alcohol-related disease, and moderate to severe renal disease, highest education achieved and socioeconomic level; Loeb et al adjusted for Charlson comorbidity index, marital status, and educational level; Lian et al adjusted for age, year of cohort entry, alcohol-related disorders, smoking status, body mass index, precancerous skin lesions, presence of naevi, immunosuppression, use of anti-parkinsonian drugs, Charlson comorbidity score, number of different drug classes used, and number of physician visits in the year before cohort entry, and health-seeking-related variables and disposable income; Matthews et al adjusted for the following number of consultations in year before index date, body mass index, alcohol use, current drinker, ex-drinker, smoking status, current smoker, and ex-smoker.). Due to these variances between studies, we should carefully interpret the results and should consider it as potential limitation. In addition, the details of melanoma were not presented in every study, which made it difficult to ascertain the results of different stages etc.Second, other confounding risk factors which are relatively critical to melanoma, such as sun exposure, have not been well evaluated. Some studies reported adjusted results, but vary from each other (e.g., Li et alThird, the majority of participants were western citizens. There are also a great amount of people who use PDE5 inhibitors in developing countries, such as China. But no study has ever explored the association in Asian or African population. Thus, the conclusion of the present study may be not applicable for other races or regions. All of the above factors may possibly affect the association strength observed in our study.5The results of the present analysis indicated that the association between PDE5 inhibitors and melanoma risk might not be causal. As for subgroup analysis, the local melanoma risk was elevated, while the nonlocalized melanoma was reduced. BCC risk was also increased, which indicated the association might not be legit. Thus, further prospective designed high-quality studies are still required to assess the association between PDE5 inhibitors and melanoma risk.


Conversely, the heterozygous LMP7 Q/K polymorphism was found to decrease the risk of ESCC in this population . Moreover, LMP2 R/C+C/C genotype was associated with increased tumor invasion depth (p = 0.041). Haplotype analysis showed that haplotype A, which includes wild-type homozygous LMP2/TAP1 and mutant LMP7, decreases susceptibility to ESCC in the Kazakh population; in contrast, haplotype E, which includes wild-type homozygous LMP2/LMP7/TAP1, acts as a risk factor for increased susceptibility to ESCC. This is the first study to report that the heterozygous LMP2 R/C and homozygous C/C genotypes increase susceptibility to ESCC in the Kazakh population and that the heterozygous LMP7 Q/K genotype decreases susceptibility to ESCC in this population. Nevertheless, neither LMP2 nor LMP7 was associated with human papillomavirus (HPV) infection. Understanding LMP2/LMP7 genetic variability will provide a new therapeutic perspective for Kazakh patients with ESCC.The Kazakh population in Xinjiang Province in northwestern China exhibits a high incidence of esophageal squamous cell carcinoma (ESCC). Although the etiology of esophageal carcinoma (EC) has not been elucidated, there are reports of the involvement of an immunologic mechanism. In the current study, 268 Kazakh ESCC patients and 500 age- and sex-matched control subjects were recruited. DNA was extracted from paraffin-embedded tumor specimens from the patients and peripheral blood lymphocytes from the controls and used for LMP2/LMP7 genotyping. Polymerase chain reaction-restriction fragment length polymorphism (PCR-RFLP) analysis was performed to detect LMP2/LMP7 gene single-nucleotide polymorphisms (SNPs). We found a clear increased risk of ESCC in the Kazakh population for the heterozygous LMP2 R/C genotype and the homozygous C/C genotype (OR = 1.470, 95%CI = 1.076–2.008,  Esophageal carcinoma (EC) is considered one of the most malignant cancers worldwide –5. More The low-molecular-weight polypeptide (LMP) system recognizes intracellular infection and plays a crucial role in immunological surveillance in humans through the MHC-I molecule and cytotoxic T-lymphocyte (CTL) pathway , 13. LMPMany studies to date have found that LMP2/LMP7 polymorphism affects susceptibility to numerous cancers. For instance, the LMP7–145 gene polymorphism contributes to increased susceptibility to gastric cancer and ovarian cancer , 24. LMPTwo hundred sixty-eight formalin-fixed and paraffin-embedded ESCC tissues were obtained from 268 Kazakh patients who had been treated with esophagectomy without any additional therapy, such as radiation and chemotherapy, during the period from January 1990 to March 2014 at People’s Hospital of Xinjiang Uygur Autonomous Region, First Affiliated Hospital of Shihezi University Medical College in Xinjiang and some hospitals in the Yili Kazakh Autonomous Region. All 268 of the cases originated from the ESCC-high-risk Kazakh population in Xinjiang. The age of the study group ranged from 23 years to 79 years, with a mean of 53.14±8.20 years. A standardized questionnaire was used to collect information on demographic characteristics  and life-style variables, including consumption of hot tea, alcohol and tobacco,we defin2, 0.5 ml each dNTP, 10 pmol primers and 1.5 U Taq Polymerase . For PCR amplification, the reaction was initially maintained at 94°C for 2 min, followed by 30 cycles of the following conditions: denaturation at 94°C for 40 s, annealing at 57.5°C for 40 s, elongation at 72°C for 40 s. A final extension at 72°C was performed for 10 min. Gene-Amp PCR System 9700 was used. The PCR products were purified using a MultiScreen-PCR purifying plate .DNA was extracted from paraffin sections and blood samples using standard proteinase K digestion and a tissue DNA extraction kit following the manufacturer’s instructions . The two previously documented polymorphisms in the LMP2/LMP7 coding regions are listed in For restriction fragment length polymorphism (RFLP) analysis, 330-bp and 351-bp PCR products for the LMP2–60 and LMP7–145 polymorphisms were digested using the specific restriction endonucleases HhaI and BsmI . The PCR products were digested under the conditions recommended by the manufacturer and then electrophoresed through a 3% agarose gel.The wild-type genotype of the LMP2–60 polymorphism yielded bands of 79 and 251 bp, and the heterozygous genotype yielded bands of 79, 251 and 330 bp . Regardi5’-CAGACACCATGGTGCACCTGAC-3’, and reverse, 5’-CCAATAGGCAGAGAGAGTCAGTG-3’. We used non-degenerate primers to determine HPV infection , which amplify a 150-bp fragment of the L1 gene in a large range of HPV types. For HPV16 detection, E7 was amplified with forward primer 5’-GATGAAATAGATGGTCCAGC-3’ and reverse primer 5’-GCTTTGTACGCACAACCGAGC-3’. For each PCR reaction, the final volume was 25 μl and included 5 μl extracted concentrated DNA. The reaction was first carried out at 95°C for 10 min, followed by 40 cycles at 94°C for 30 s, 42°C for 90 s, 72°C for 30 s, and a final extension at 72°C for 5 min. The amplicons were then denatured and subjected to hybridization. The samples were tested in triplicate, and β-actin was used as an internal control.DNA was extracted from 268 paraffin-embedded tissue samples from Kazakh patients using standard proteinase K digestion and a tissue DNA extraction kit . We used a human β-actin primer set as an internal control for PCR: forward, p values were corrected based on the number of specimens tested and the number of comparisons performed. Student’s t-test and the χ2 test were utilized to calculate the significance of differences in continuous variables between LMP genotypes. The relationship between single-nucleotide polymorphisms (SNPs) in LMP2/LMP7 and the risk of ESCC was estimated by determining the odds ratios (ORs) at 95% confidence intervals (CIs) using logistic regression analyses for crude ORs and adjusted ORs when adjusting for age and sex. Logistic regression was also applied to analyze the HPV status and risk of ESCC.All statistical analyses were performed using SPSS version 17.0 (U.S.). Hardy-Weinberg equilibrium (HWE) for the cases and controls was evaluated using chi-square tests to compare the observed and expected genotype frequencies . Allele p = 0.095 and p = 0.639, respectively). Statistical analysis showed that smoking is closely linked to the risk of ESCC in the Kazakh population (p = 0.027), though drinking was not significantly associated (p = 0.393). We also evaluated data for the case group with regard to differentiation status, tumor location, depth of invasion and lymph node metastasis.The characteristics of the 268 cases and 500 controls included in this study are summarized in p = 0.015 for R/C, OR = 2.048, 95%CI = 1.168–3.591, p = 0.011 for C/C, p = 0.003). In the dominant model, the RC+CC genotype was associated with a statistically increased risk of ESCC in the Kazakh population when the RR genotype was used as a reference group . The dominant model, for which the QQ genotype was used as a reference group, showed that the QK+KK genotype was associated with a statistically increased risk of ESCC in the Kazakh population , p = 0.259), https://gtexportal.org/home/) to study the effects of SNP locus of LMP2/LMP7 gene on gene expression, the results showed that SNP locus of LMP2 in esophageal tissues affected the expression of LMP2 , 14.9% (Q/K), and 3.73% (K/K) among the cases and 68.4% (Q/Q), 29.8% (Q/K), and 1.8% (K/K) among the controls. Compared with the Q/Q genotype, the Q/K genotype decreased the risk of ESCC in the Kazakh population . There was no statistically significant with drinker and nondrinker(p>0.05). Equally contrary, the LMP7–145 Q/K and QK+KK genotypes were associated with a decreased risk of ESCC with smoking people but not in those without smoking(p<0.05). Nevertheless, the result indicated that LMP7–145 Q/K and QK+KK genotypes were associated with a decreased risk of ESCC who don't drink but not in those with drinking(p<0.05). When Kazakh ESCC patients were divided into two subgroups, T1/T2 and T3/T4, according to the AJCC TNM classification of EC, the RC+CC genotype was significantly associated with ESCC tumor depth . However, there was no significant association between LMP2/LMP7 polymorphism and ESCC in the Kazakh population with respect to clinical pathological parameters (p>0.05).To evaluate the contribution of confounding factors, including gender /age/ smoking and drinking to the risk of ESCC in the Kazakh population, stratification analyses were performed to investigate the potential effect of genetic variants of LMP2/LMP7 with ESCC risk in population subgroups . This anp<0.001), whereas haplotype E, which includes wild-type homozygous LMP2/LMP7/TAP1, acts as a risk factor to increase susceptibility to ESCC in the Kazakh population . The results are summarized in The association between TAP1 polymorphisms and ESCC in the Kazakh population was shown in a previous study (40). Thus, haplotype analysis was performed to further evaluate the combined effects of LMP2/LMP7 and TAP1 polymorphisms on the risk of ESCC in this population. Among the eight haplotypes derived from the observed genotypes, A and E were the most common in both the cases and controls  . The resp>0.05).The results of genotype frequency analysis for LMP2/LMP7 gene polymorphisms in HPV-positive and HPV-negative patients are summarized in ESCC is a polygenic hereditary disease that can be affected by genetic factors –30. In oAccording to our results, heterozygous LMP2 R/C and homozygous LMP2 C/C genotypes are risk factors for ESCC in the Kazakh population. This finding differed from that of Xiang Ma and BangHowever, one study did find that LMP2 plays a role in the growth of multiple myeloma and acute myeloid leukemia but that LMP7 is not associated with these hematological malignancies . NeverthIn subgroup analysis, the LMP2 R/C genotype was associated with increased tumor invasion depth in ESCC, suggesting that genetic variation in immune response genes may play a role in ESCC progression. Additionally, no association between LMP7–145 polymorphism genotypes and ESCC clinicopathological characteristics of the Kazakh population was observed in our study. Song  revealedThe LMP2–60 C/C and R/C+C/C genotypes were associated with an increased risk of ESCC in Kazakh patients aged ≥57 years but not in those aged <57 years. This indicates that carcinogenesis is an accumulation of genetic events during aging . AdditioOur previous studies found that TAP1 polymorphisms could increase susceptibility to ESCC in the Kazakh population . MoreoveTo date, many studies have confirmed that HPV is present in ESCC samples, with infection rates ranging from 17.1  to 78.11In conclusion, to the best of our knowledge, the present study is the first to demonstrate that LMP2/LMP7 polymorphisms are associated with the risk of ESCC in the Kazakh population, a Chinese ethnic minority. LMP2 can be considered a risk factor for ESCC and LMP7 as a protective factor against ESCC in the Kazakh population. Additional studies with larger sample sizes and the inclusion of various populations, as well as functional studies, are required to verify the present preliminary findings.S1 Fig(PDF)Click here for additional data file.S2 Fig(PDF)Click here for additional data file.S3 Fig(PDF)Click here for additional data file.S1 Table(PDF)Click here for additional data file.S2 Table(PDF)Click here for additional data file.S3 Table(PDF)Click here for additional data file.


Diabetes is a major cause of morbidity for patients undergoing surgery and can increase the incidence of some postoperative complications such as bedsores. We conducted a meta-analysis of observational studies to examine whether patients with diabetes undergoing surgery had high risk of bedsore. We performed a systematic literature search in Pubmed, Embase and the Cochrane Library Central Register of Controlled Trials database from inception to November 2016. Studies were selected if they reported estimates of the relative risk (RR) for bedsore risk in postoperative diabetic patients compared with that of in non-diabetic patients. Random-effects meta-analysis was conducted to pool the estimates. A total of 16 studies with 24,112 individuals were included in our meta-analysis. The pooled RR of bedsore development for patients with diatetes was 1.77 (95% CI 1.45 to 2.16). The results of subgroup analyses were consistent when stratified by surgery type, study design, research region, sample size, inclusion period, analysis method and study quality. There was evidence of publication bias among studies and a sensitivity analysis using the Duval and Tweedie “trim-and-fill” method did not significantly alter the pooled results .This meta-analysis provides indications that diabetic patients undergoing surgery could have a higher risk of developing bedsores. Further large-scale prospective trials should be implemented to comfirm the association. Bedsore, also known as pressure ulcer, is a common cause of prolonging length of hospital stay for patients with surgery. It has been reported that the length of hospital stay of surgical patients could increase by 3.5 to 5 days on average when a bedsore occurs , 2. For Several risk factors and aetiologies have been reported to contribute to the development of bedsores during perioperative period. Traditionally, it is considered that patients with advanced age, malnutrition , poor circulation or smoking may have a higher risk of bedsores –7. MoreoNumerous studies have explored the role of patients with preexisting diabetes on the development of bedsore. Despite the fact that some studies have reported significant association between diabetes and risk of surgery-related bedsore, some others have reported varying results on this association. It was noted in several studies that surgical patients with diabetes had higher risk of bedsore than those without diabetes –18, whilThe initial literature search yielded 1046 abstracts of which 31 were considered potentially relevant for full-text review. Totally, 16 studies including 24,112 participants met our eligibility criteria and were involved in the meta-analysis , 26–29. Table 2 = 62.7%; P < 0.001)  and there was statistical inter-study heterogeneity (IP = 0.009).Table P = 0.238) was noted.Four types of sugery were involved in the studies, with 6 of general surgery, 4 of hip surgery, 4 of cardiac surgery and 2 of lower extremity amputations, respectively. The summary RRs estimated for bedsore incidence were 1.71 (95% CI 1.40 to 2.09) for general surgery, 1.78 (95% CI 1.14 to 2.78 ) for hip surgery, 1.98 (95% CI 1.41 to 2.79) for cardiac surgery and 1.44 (95% CI 0.93 to 2.24) for lower extremity amputations, respectively. No statistically significant difference for inter-study heterogeneity (P = 0.017).As demonstrated in Table P = 0.019).The summarised RRs for bedsore risk stratified by sample size were 1.66 (95% CI 1.21 to 2.29) for studies with large sample size (≥ 1000) and 1.93 (95% CI 1.57 to 2.38) for studies with small sample size (< 1000). We found statistically significant difference for inter-study heterogeneity (P = 0.523).Six and 9 studies were conducted in Europe and USA, respectively. The summary RRs for bedsore risk were 1.94 (95% CI 1.26 to 2.99) for studies conducted in Europe and 1.62 (95% CI 1.33 to 1.97) for studies conducted in USA. No statistically significant difference for inter-study heterogeneity was found (P = 0.089).Four studies included participants before year 2000 and the pooled RR was 1.38 (95% CI 1.09 to 1.76); and 4 included participants after year 2000 with the pooled RR of 1.61 (95% CI 1.30 to 2.00). We did not find statistically significant difference for inter-study heterogeneity (P = 0.078).The summarised RRs for bedsore risk stratified by patient age were 1.66 (95% CI 1.19 to 2.32) for studies with patients having older age (≥ 70 years) and 1.77 (95% CI 1.48 to 2.11) for studies with patients having less older age (< 70). There was no statistically significant difference for inter-study heterogeneity (P < 0.001) was found.Ten studies applied univariate analysis to analyze the risk estimates and 6 studies applied multivariate analysis. The results showed that the pooled RRs for bedsore risk were 2.08 (95% CI 1.73 to 2.50) for studies with univariate analysis and 1.44 (95% CI 1.11 to 1.88) for studies with multivariate analysis. Statistically significant difference for inter-study heterogeneity (P = 0.002). We further test whether publication bias significantly influenced the pooled risk estimates by using trim and filled method and the adjusted RR indicated the same trend with the result of the primary analysis (RR 1.17 95% CI 1.02 to 1.36). A sensitivity analysis was carried out by excluding one study at each time and then recalculating the pooled RRs for the remaining ones to test the effect of each study on the overall estimates. We did not find the alteration of in the direction of the estimate when any one of the included study was excluded. This analysis confirmed the robustness of the positive association between diabetes and bedsore risk in surgical patients.The shape of the funnel plot for the studies on the diabetes and bedsore risk seemed asymmetrical. In addition, Egger's adjusted rank correlation test showed potential evidence of publication bias , though no statistical significance is found among different surgery types in this meta-analysis.This updated meta-analysis further confirms and extends the preliminary findings of the two previous published meta-analyses , 25. TheMultiple mechanisms can contribute to the deveopment and severity of bedsores, which may result from capillary occlusion by external pressure, leading to the shut off of blood supply, cell death, necrosis removal and ulceration. The severity of the bedsore is determined by the length of time pressure is applied to the local region. Moreover, for a patient receiving surgery, the incidence rate of bedsores is mainly determined by the duration and intensity of the shearing force given upon the tissue during surgery. For the impact of different surgery types on the risk of bedsore development, we noted that patients with cardiact surgery had the higher risk  than patients with general surgery or hip surgery, while patients with lower extremity amputations had the lowest risk . We propose that the trauma severity of surgery to the body could be a major influential factor determining the risk of developing bedsore.2 = 62.7%, Pheterogeneity < 0.001). Sensitivity analyses indicated that exclusion of any one of the study did not significantly alter the summary estimate. The trim-and-fill model and multiple subgroup analyses stratified by some main clinical variables were in agreement with the initial findings, indicating that the result of this meta-analysis was robust and not affected by publication bias. Nevertheless, we should interpret the results with caution due to the common occurance of publication bias . Tw. Tw31]. Two reviewers (M.L. and Q.C.) independently assessed all records through reading the titles and/or abstracts for potentially eligible studies. In case there were different opinions, a senior reviewer (L.L ) would join to discuss and resolve the disagreement. We included studies in this meta-analysis if they satisfied the following criteria: (i) observational studies including cohort or case–control studies; (ii) investigating diabetes and risk of bedsore in patients undergoing surgery; (iii) providing odds ratios (ORs)/relative risks (RRs) along with 95% confidence intervals (95% CIs) or sufficient information to calculate them, for bedsore risk stratified by diabetes in patients having operation. We included patients with history or diagnosis of diabetes, irrespective of diabetes type (1 or 2), disease severity, duration or anti-diabetic drug use due to unavailability of those data.Data were extracted independently according to a predesigned form by two reviewers (Y.Z. and L.H.) and the results were crosschecked. A third reviewer(L.L.) would reevaluated the extracted data if any disagreements occurred. The following data were extracted from each study: first author, publication year, study region, study design, inclusion period, opertion site, number of participants, sex, mean/median age, body mass index, treatment regimen, analysis method, follow up period, adjustment variables, and risk estimates for association between bedsore risk and diabetes in patients having operation.Two reviewers (M.L. and L.H.) independently assessed the methodological quality of each included study using Newcastle-Ottawa quality assessmentscale (NOS) for cohort studies, which included 3 domains  totaling 9 points (Table 2, with an I2 statistic more than 50% defining significant heterogeneity [P value less than 0.05 indicating significant difference.We quantified the relationship between diabetes and risk of bedsore using an inverse variance method using DerSimonian and Laird random-effects models . All staogeneity , 34. We ogeneity . If publogeneity . Sensitiogeneity . All sta


P = 0.01; CT + TT vs. CC: OR = 1.21, 95% CI = 1.04–1.41, P = 0.01) in the total population, as well as in Caucasian subgroup. The 2677G > T/A SNP was related to a decreased AD risk in Caucasian subgroup . Moreover, the ABCB1 haplotype analysis showed that the 1236T/2677T/3435C haplotype was associated with a higher risk of AD . Our results suggest that the ABCB1 3435C > T SNP, the 2677G > T/A SNP and 1236T/2677T/3435C haplotype are significantly associated with AD susceptibility.Although several epidemiological studies have investigated the association between ATP-binding cassette subfamily B member 1 (ABCB1) gene polymorphisms and Alzheimer’s disease (AD) susceptibility, controversial results exist. Here, we performed a meta-analysis to assess whether ABCB1 polymorphisms 3435C > T (rs1045642), 2677G > T/A (rs2032582), 1236C > T (rs1128503) and haplotypes were associated with AD risk. Nine independent publications were included and analyzed. Crude odds ratio (OR) and 95% confidence interval (CI) were applied to investigate the strength of the association. Sensitivity analysis was conducted to measure the robustness of our analysis. A funnel plot and trim and fill method were used to test and adjust for publication bias. The results showed a significant association between the 3435C > T single nucleotide polymorphism (SNP) and AD susceptibility (CT vs. CC: OR = 1.24, 95% CI = 1.06–1.45,  Alzheimer’s disease (AD) is the most common progressive neurodegenerative disorder of the central nervous system, clinically characterized by memory impairment and cognitive dysfunction. The pathological hallmarks of AD in the brain are massive cerebral accumulation of senile plaques largely composed of β-amyloid (Aβ) peptide, intracellular neurofibrillary tangles and neuronal loss12378910The ATP-binding cassette subfamily B member 1 (ABCB1) gene in the chromosome 7q21 region spans 209-kb and contains 29 exons1314151719et al.et al.23242627et al.P-glycoprotein (P-gp) is encoded by the ABCB1 gene and acts as an integral component of the blood-brain barrier (BBB)23242526272829302425272930P > 0.05). The key information of the studies on ABCB1 3435C > T, 2677G > T/A, 1236C > T polymorphisms and AD risk is summarized in Fourty-nine potential studies were identified after an initial search from the Pubmed, Embase, Web of Science, Medline and Alzgene databases. After further screening, nine studies were finally enrolled in our meta-analysis (2366 cases and 2256 controls). The flow diagram of the search process was shown in P = 0.82; CT vs. CC: OR  =  1.18, 95% CI = 0.97–1.45, P = 0.10; CT + TT vs. CC: OR = 1.14, 95% CI = 0.91–1.42, P = 0.26; TT vs. CT + CC: OR = 0.91, 95% CI = 0.75–1.11, P = 0.36; T vs. C: OR = 1.01, 95% CI = 0.87–1.17, P = 0.90. A moderate heterogeneity was found between the individual studies . We conducted a sensitivity analysis by excluding each study, and found the Fehér’s study2 decreased to 0% with a different but statistically significant pooled estimate in the homozygote comparison model  and corresponding 95% confidence intervals (CIs) in all the genetic models were as follows: TT vs. CC: OR = 1.04, 95% CI = 0.76–1.41, P = 0.03, P = 0.04, Additionally, subgroup analysis by ethnicity showed a significant association between 3435C > T SNP and an increased risk of AD among Caucasians . However, the funnel plots after the adoption of trim and fill method had perfect symmetry and no significant difference was found between the ORs and 95% CIs before and after, suggesting that there were no change within the crude and adjusted estimates . An obvious heterogeneity was observed in most models . However, subgroup analysis according to the ethnicity in the recessive model showed a significant association between 2677G > T/A SNP and a decreased AD risk among Caucasians  with no heterogeneity , but not in mixed , as well as the Egger’s linear regression test , suggesting that there was no publication bias in this meta-analysis. Pooled ORs and 95% CIs for the association of ABCB1 2677G > T/A SNP with AD risk are shown in Five studies investigated the association of 2677G > T/A SNP with the AD risk. No significant association between 2677G > T/A SNP and AD susceptibility was detected in the total populations . The TTT and CTT haplotypes exhibited no association with the AD risk  haplotype, the 1236T/2677T/3435C (TTC) haplotype and the 1236C/2677T/3435T (CTT) haplotype . The pooAs one of the most common forms of dementia, AD affects millions of people worldwide. Recently, several studies have revealed that mutations of ABCB1 can be responsible for the pathogenesis of AD by promoting the accumulation of Aβ via P-gp22232423242527282930et al.et al.Fehér et al.Consistent with the report by Cascorbi et al.et al.et al.et al.2324Due to the existence of multiple genes in linkage disequilibrium, haplotype analysis was useful to identify other indirect SNPs at the same region. As for the ABCB1 gene, the TTT, TTC and CTT haplotypes were most investigated for their roles in AD susceptibility. The studies by Fehér et al.A recent Genome-Wide Association Study (GWAS) by Lambert There are some limitations in this meta-analysis. First, although we collected all the eligible publications, studies included in our meta-analysis were still limited. Randomized controlled clinical trials with larger sample sizes are required to confirm our conclusion. Second, AD is a multifactorial disease, and the potential interactions among gene-gene and gene-environment should take into account. Many other factors such as ApoEε4 may participate in the association between ABCB1 polymorphisms and AD risk. However, we did not carry out a subgroup analysis based on it due to limited data. Third, since some studies have not provided informations such as age, gender, diagnostic criteria and sequencing method, we could not carry out a subgroup analysis to study the effects of these parameters on the association between ABCB1 polymorphisms and AD risk. To better understand the role of ABCB1 polymorphisms or haplotypes in AD susceptibility, large-scale, high-quality studies with multi-ethnic populations are needed.In conclusion, based on the published studies, our meta-analysis confirmed that the ABCB1 3435C > T SNP and TTC haplotype were significantly associated with an increased AD risk, while the 2677G > T/A SNP may contribute to a lower susceptibility of AD. Our study, for the first time, found that there were associations between ABCB1 polymorphisms and haplotypes and AD risk. Our findings provide a better understanding of AD pathogenesis, and ABCB1 polymorphisms and haplotypes may serve as important targets for the prevention and diagnosis for AD.To identify studies eligible for the meta-analysis, a comprehensive search strategy was applied using the electronic databases including Pubmed, Embase, Web of Science, Medline and Alzgene. The following keywords were used: ABCB1 or ATP-binding cassette subfamily B member 1, and polymorphism or rs1045642 or rs2032582 or rs1128503, and Alzheimer’s disease or AD. All selected articles were published from January 2001 to March 2016. Other relevant studies were identified by hand searching the references of included articles.Studies included in this meta-analysis met the following criteria: (1) the study had to be focused on the 3435C > T, 2677G > T/A, 1236C > T SNP or haplotypes within the ABCB1 gene using original data; (2) the study should be case-control-designed or cohort-designed; (3) all patients must meet the diagnostic criteria of AD; (4) both AD patients and controls were included; and (5) the frequencies of individual genotypes in cases and controls were reported or provided from the authors. Exclusion criteria included studies with no relevance to ABCB1 polymorphisms and AD risk, overlapping data, and review articles. Each article was checked with inclusion and exclusion criteria independently by two investigators (X.Z. and X.-H.S.).P < 0.05 was considered as statistically significant. We adopted the random-effects model to calculate the pooled ORs when a significant heterogeneity existed in the genetic models in the initial overall analysis, otherwise a fixed-effects model should be applied. The between-study heterogeneity was estimated using Cochran’s Q statistic and the I2 statistic. P > 0.10 in Q-test2 < 25%P < 0.05To assess the association between ABCB1 3435C > T, 2677G > T/A, 1236C > T polymorphisms and AD risk, ORs and corresponding 95% CIs were calculated using homozygote comparison (BB vs. AA), heterozygous comparison (AB vs. AA), dominant model (BB + AB vs. AA), recessive model (BB vs. AB + AA), and allele comparison (B vs. A) . The significance of the pooled OR was determined by Z-test, and How to cite this article: Zhong, X. et al. Association between ABCB1 polymorphisms and haplotypes and Alzheimer’s disease: a meta-analysis. Sci. Rep. 6, 32708; doi: 10.1038/srep32708 (2016).