Background: The Arg213His (rs9282861) polymorphism of Sulfotransferase Family 1A Member 1(SULT1A1) gene has been reported to be associated with risk of breast cancer in some epidemiologicalstudies. However, the results of these studies are conflicting and inconclusive. Therefore, thissystematic review and meta-analysis was conducted to evaluate the association of SULT1A1 Arg213His(rs9282861) polymorphism with susceptibility to breast cancer.Study design: A systematic review and meta-analysis.Methods: A comprehensive literature search for eligible studies was conducted in PubMed, Elsevier,Science Direct, Scopus and Google Scholar databases up to October 5, 2017. Pooled odds ratios (ORs)with their corresponding 95% confidence intervals (95% CIs) were used to evaluate the strength of theassociation using fixed effects models and random effects models.Results: A total of 20 relevant case-control studies involving 11,077 cases and 14,798 controls wereincluded in this meta-analysis. Overall, there was a significant association between the SULT1A1Arg213His (rs9282861) polymorphism and risk of breast cancer in the allele mode  and the homozygote model . Subgroup analysis based on ethnicity suggested SULT1A1 Arg213His (rs9282861)polymorphism had a subtly increased breast cancer risk among Asian population, but not Caucasians.Further, subgroup analyses, significant associations were observed in hospital based group, RFLP-PCRgroup, and high quality studies subgroups.Conclusions: This meta-analysis suggested that SULT1A1 Arg213His (rs9282861) polymorphismmight be associated with breast cancer risk, especially among Asian population. Moreover, theSULT1A1 Arg213His polymorphism is of high clinical relevance by ethnicity and would be a usefulmarker to identify patients who are at higher risk for breast cancer. Currently, more women survive due to earlier diagnosis and better therapy 4. However, the morbidity and mortality of breast cancer have increased in most developing countries5. It is well-established that family or personal history of cancer, nulliparous, and history of hormone replacement therapy and different genetic background are most known risk factors for breast cancer 6. Breast cancer is one of the first multifactorial conditions to be evaluated using molecular genetics techniques 8. It is estimated that 15- 30% of hereditary breast cancer cases are due to known highly penetrant genes such as BRCA1 and BRCA2 9. More than 1,000 different breast cancer-associated mutations in each gene have been documented. However, a few of these mutations could be clearly identified as pathological 11.Breast cancer is one of the most common malignancies among women worldwide SULT1A1 gene are increasingly studied for an increased different cancer risk because of the critical roles in catalyzing the sulfate conjugation of many hormones, neurotransmitters, drugs, and xenobiotic compounds 12. In particular, SULT1A1 has high activity towards a wide range of substrates including environmental carcinogens such as tobacco. The SULT1A1 gene is located on chromosome 16p12.1-p11.2, spans approximately 5.1 kb and contains nine exons that range in length from 74 to 347 bp 13. To date, the Arg213His (rs9282861) polymorphism is most widely studied SNP within SULT1A1 gene. The Arg213His polymorphism is located in exon 7 with G to A transition, at nucleotide 638 (codon 213) leading to an arginine (Arg) to histidine (His) amino acid substitution 14. Functional studies have reported that the variant A allele  is associated with lower sulfotransferase activity and thermal stability compared with the wild-type G allele  16.Genetic variations in SULT1A1 Arg213His (rs9282861) polymorphism with breast cancer risk 35. On the whole, these studies results were conflicting and inconclusive, probably due to the relatively small size, ethnic background, uncorrected multiple hypothesis testing, and publication bias. It is known that meta-analysis is a statistical procedure for combining data from multiple studies to produce a single estimate on the same outcome of interest with enhanced evaluation. Therefore, current systematic review and meta-analysis was performed from all eligible studies to derive a more precise estimation the association between SULT1A1 Arg213His (rs9282861) polymorphism and breast cancer risk.So far, there were so many reports about the association of SULT1A1 Arg213His (rs9282861) polymorphism and the risk of breast cancer up to October 5, 2017. The search strategy identified all possible studies using combinations of the following keywords: Sulfotransferase Family 1A Member, SULT1A1 gene, G638A, Arg213His, rs9282861, and polymorphism, genetic polymorphism, single nucleotide polymorphism, SNP mutation, and breast, breast cancer, breast malignancy, breast tumor, breast carcinoma, and breast neoplasm. In addition, manual searching was carried out to ensure that no relevant studies were missed in the database search by scanning the review articles, clinical trials, and meta-analyses. No language restrictions were imposed.We performed a systematic literature search of electronic databases such as PubMed, Embase, ISI Wed of Knowledge, Google scholar, Scopus, Cochrane Library, Chinese National Knowledge Infrastructure (CNKI) and Chinese Wan Fang databases to identify studies that evaluated the association between SULT1A1 Arg213His (rs9282861) polymorphism and breast cancer risk; c) providing sufficient data to estimate the odds ratio (OR) with the corresponding 95% confidence interval (95%CI). Studies were excluded if the following criteria were satisfied: a) Abstracts, case reports, case-only articles, family-based studies, reviews articles and repeated literatures; b) duplicate of previous publications; c) not provided sufficient data (e.g. neither the frequency nor the number of genotype was given) for calculation.To be eligible for the inclusion criteria in the meta-analysis, the following criteria were used: a) case-control or cohort design; b) evaluating the associations between P value of the Hardy–Weinberg equilibrium (HWE) test in the control. Different ethnic descents were classified as Caucasian, Asian, Latinos or Mixed (derived from an admixture of different ethnic groups). Disagreements between the two investigators were resolved by discussing with a third investigator. If more than one article were published using the same data, only the newest study and with largest sample size was selected. Moreover, the quality of selected studies was tested by the confirmation of HWE in control groups, and studies without the confirmation of HWE in controls were defined as low-quality studies, while studies with the confirmation of HWE in controls were defined as high-quality studies.All data were extracted independently by two authors using a standardized data extraction form and according to the inclusion criteria listed above. The following elements were extracted for each included study: first author name, year of publication, country, ethnicity, source of control groups, genotyping method, total number of cases and controls, minor allele frequency (MAF) in control subjects, SULT1A1 Arg213His (rs9282861) polymorphism and breast cancer risk. The pooled ORs were determined for SULT1A1 Arg213His (rs9282861) polymorphism under the allele model (A vs. G), the homozygote model (AA vs. GG), the heterozygote model (AG vs. GG), the dominant model (AA+AG vs. GG), and the recessive model (AA vs. AG+GG). The significance of the pooled OR was determined by the Z-test, in which P<0.05 was considered as statistically significant. The Chi-square-based Q-test and I2-statistics were employed for evaluating the between-study heterogeneity36,37. A P-value greater than 0.10 for the Q-test or I2 less than 50% indicates a lack of heterogeneity among studies, so the pooled OR estimate of the included studies was obtained by the fixed-effects model (Mantel-Haenszel method) 38. Otherwise, the random-effects model (DerSimonian-Laird method) was used 39. Hardy-Weinberg equilibrium (HWE) in the control group was calculated by the Chi-square test for goodness of fit, If P value > 0.05, the genotype distribution of control population conformed to HWE. Furthermore, to explore the sources of heterogeneity, subgroup analysis we conducted based on ethnicity, source of controls and studies quality. To validate the reliability of the results, sensitivity analysis was performed though omitting one case-control study each time, as well as limiting this meta-analysis to studies which were conformed to HWE. Publication bias was evaluated by Begg’s test and Egger’s test 40,41. If publication bias existed, the Duval and Tweedie non-parametric ‘‘trim and fill’’ method was used to adjust for it. All statistical analyses including meta-regression, sensitivity analysis, subgroup analysis, and publication bias assessment were performed using Comprehensive Meta-Analysis (CMA) software version 2.0 . Two-sided P values < 0.05 were considered statistically significant.Crude odds ratios (ORs) with corresponding 95% confidence intervals (CIs) were used to assess the strength of association between the SULT1A1 gene, not related to breast cancer or reported other diseases. Finally, a total of 20 studies with 11,077 cancer cases and 14,798 controls met the inclusion criteria. Study characteristics are summarized in SULT1A1 Arg213His (rs9282861) polymorphism was performed by five genotyping methods including real-time PCR (RT-PCR), PCR-restriction fragment length polymorphism (PCR-RFLP), TaqMan, Genotyping by matrix-assisted laser desorption/ionization time-of-flight mass spectrometry (MALDI-TOF MS) and direct Sequencing. The distribution of genotype in the controls of the studies was not in agreement with Hardy-Weinberg equilibrium in six studies, which were further tested in the sensitivity analyses and subgroup analysis of studies with high quality. The detailed characteristics of each study and SULT1A1 Arg213His (rs9282861) polymorphism genotype distributions included in the meta-analysis are presented in As shown in SULT1A1 Arg213His (rs9282861) polymorphism and breast cancer risk. The random effects model was chosen to synthesize the data from the allelic model, homozygote, heterozygote, dominant model and recessive model. Overall, there was a significant association between SULT1A1 Arg213His (rs9282861) polymorphism and breast cancer in the allele mode  and the homozygote model , but not in the heterozygote model , the dominant model , and the recessive model .SULT1A1 Arg213His (rs9282861) polymorphism and breast cancer was observed in Asians in the allele model , the homozygote model , the heterozygote model , the dominant model  and the recessive model , but not in the Caucasian populations.In the subgroup analysis by the ethnicity, a significant association between the SULT1A1 Arg213His polymorphism and risk of breast cancer were found in the allele mode , the homozygote model , and in the heterozygote model . Interestingly, the stratifying by source of population showed that SULT1A1 Arg213His (rs9282861) polymorphism was significantly associated with breast cancer risk in the allele mode  and the homozygote model , the dominant model , and the recessive model  in the hospital based group, but not population based (SULT1A1 Arg213His (rs9282861) polymorphism and increased risk of breast cancer in the allele model , the homozygote model  and the dominant model . In the PCR-RFLP group, significantly increased association between on based . SubgrouSULT1A1 Arg213His (rs9282861) polymorphism in the healthy controls is presented in SULT1A1 Arg213His (rs9282861) polymorphism exhibited ethnic variations. The allele and genotype distributions of 213His allele in the Caucasian and Asians populations were 32.75% (17.5%-48.0%) and 7.4% (2.1%-12.7%), respectively. Thus, the frequency of the 213His allele in Asians was less than Caucasians.The MAFs for the SULT1A1 Arg213His (rs9282861) polymorphism , P= ≤0.001, I2 = 73.77%; and dominant model: Q= 189.01 on 17 d.f (Q), P= ≤0.001, I2 = 91%). Therefore, we have performed subgroup analyses by ethnicity and HWE status to explore the source of heterogeneity. The I2 decreased obviously and P value exceeded 0.05 under the homozygote, heterozygote and recessive models among Asians, while it was still significant under all genetic models among Caucasians  polymorphism in the allele mode (A vs. G: P<0.001), the homozygote model (AA vs. GG: P<0.001), the heterozygote model (AG vs. GG: P=0.004), the dominant model (AA+AG vs. GG: P<0.001), and the recessive model (AA vs. AG+GG: P=0.008). Then, the source of heterogeneity was assessed by meta-regression analysis. However, the ethnicity, genotyping methods, studies quality and source of population did not contribute to substantial heterogeneity in this meta-analysis  polymorphism with risk of breast cancer in dominant model  polymorphism and breast cancer. The pooled ORs of 20 case-control studies with 11,077 cases and 14,798 controls in the current meta-analysis concluded that there was a significant correlation between SULT1A1 Arg213His (rs9282861) polymorphism and breast cancer under allele mode  and homozygote model .The pathogenesis of breast cancer is complex and genetic factor play an important role in breast cancer susceptibility. An increasing number of publications on genetic association studies, genome-wide association studies (GWASs), and relate meta-analyses have been published to clarify the association between different gene polymorphisms and breast cancer. To our knowledge, this is the most comprehensive and robust meta-analysis to explore and report the association between the SULT1A1 Arg213His polymorphism and breast cancer risk 12. The current meta-analysis is different from previous meta-analysis. We collected all 20 eligible studies, which include 13 studies used by Xiao et al., researches and seven new cases-control studies. It was remarkable that they have not included the Choi et al. 2003 22, Le Marchand et al. 2005 24, Yang et al. 2005 27, Han et al. 200629, and Gulyaeva et al. 2008 30 studies with that published up to their electronic researches due to lack of agreement with HWE. We analyzed the association between SULT1A1 Arg213His (rs9282861) polymorphism and breast cancer in Caucasian and Asian populations. The association is significant in Asians under in the allele model , the homozygote model , the heterozygote model , the dominant model  and the recessive model , but not among Caucasian populations. Our findings suggested that SULT1A1 Arg213His (rs9282861) polymorphism might have a susceptible nature in the Asians, but not in the Caucasians. Therefore, different populations lived in different environments and gene-environment interactions may partly affect the breast cancer susceptibility. Compared with the current meta-analysis, Xiao et al. not performed subgroup analysis by ethnicity for breast cancer 12. Therefore, Xiao et al., study essentially remains an open field as meta-analysis their results reliability due lack of subgroup analysis by ethnicity and the number of studies were considerably smaller than that needed to receive the robust conclusions.Our results are inconsistent with the previous meta-analysis by Xiao et al., which not reported a significant association between SULT1A1 Arg213His polymorphism and increased risk of breast cancer in participants of those studies involving PCR-RFLP under the allele mode , the homozygote model , and in the heterozygote model . We suggested that this trend is possible for that most of the included studies mainly utilized PCR-RFLP method to genotyping. Moreover, this result should be carefully interpreted and confirmed by conducting further analysis of additional published studies.Interestingly, stratified analysis according to genotyping method showed a significant association between 44. The study designs, source of controls subjects, ethnicity, genotyping method, sample size, lifestyle, subject’s age and gender may contribute to the heterogeneity 45. In the present meta-analysis, significant between-study heterogeneity was detected across studies under all models; therefore, we utilized the random-effects model to summarize the ORs. Additionally, in order to make the conclusion more credible, we performed the publication bias analysis and sensitivity analysis. Funnel plots suggested that no obvious publication bias was observed. Xiao et al. 12 removed the study conducted by Khvostova et al. 34 due to the low-quality and results indicated that this study influences the pooled estimates and the heterogeneity most in breast cancer subgroup. We have also removed Khvostova et al. 34 study, but there was no obvious effect on the pooled ORs and the corresponding CIs. In addition, we have performed the sensitivity analysis by removing those studies departed from HWE, but the results did not change meaningfully by excluding six studies departed from HWE or one study with low-quality. We suggest the relative small sample size status may be the reason for Xiao et al. 12 observations.Between-study heterogeneity might distort the conclusion of a meta-analysis SULT1A1 Arg213His (rs9282861) polymorphism are limited for the subgroup analysis, since there were not enough studies in other population or regions such as Africans and Middle East in this meta-analysis. Therefore there was insufficient statistical power to demonstrate the associations between the  SULT1A1 Arg213His (rs9282861) polymorphism and breast cancer risk in Africans or Middle East. Second, we only included articles that were published in English, and language bias might exist. Third, because of moderate to high heterogeneity in all five genetic models, we used the random effect model, which is not as reliable as the fixed-effects model and may potentially restrict the interpretation of the pooled risk estimates. Although the degree of heterogeneity was reduced (but not disappeared) by subgroup analyses based on ethnicity and HWE status, other sources of heterogeneity were not checked. Forth, in the current meta-analysis the design of all included studies was case control, therefore results more prone to selection bias. Fifth, the current meta-analysis results were based on unadjusted estimates, while a more precise analysis should be performed if all individual data available, which would allow for the adjustment by other co-variants including age, environmental exposures, smoking status, reproductive history, menopausal status and other lifestyle factors. Finally, gene-gene and gene-environment interactions which may modulate the breast cancer susceptibility were limited owing to the lack of the sufficient data in the eligible studies.Despite the advantage of large sample size and no evidence of publication bias, the meta-analysis had several limitations that should be taken into account. First, the population data of SULT1A1 Arg213His (rs9282861) polymorphism and risk of breast cancer. The present meta-analysis provided evidence of association between SULT1A1 Arg213His (rs9282861) polymorphism and increased breast cancer risk, especially among Asians. These results suggest that the SULT1A1 Arg213His (rs9282861) polymorphism is of high clinical relevance among Asians and would be a useful marker to identify women who are at higher risk for breast cancer. Moreover, well-designed epidemiological studies with larger sample size, carefully matching cases and control subjects and more ethnic groups are needed to validate these results. Further studies may focus on the influence of gene-gene and gene-environment interactions on the association of SULT1A1 Arg213His (rs9282861) polymorphism and breast cancer.In summary, despite the limitations, this study is the most comprehensive, best available evidence and accurate meta-analysis focusing on the association between We thank Prof. Seyed Mehdi Kalantar and Prof. Mohammad Hassan Sheikhha for their great support acknowledged.The authors declared no conflict of interest.This research received no specific grant from any funding agency in the institute or university. SULT1A1 Arg213His polymorphism is associated with increased risk of breast cancer.The  SULT1A1 Arg213His polymorphism may be associated with increased risk of breast cancer in Asians.The  SULT1A1 Arg213His polymorphism would be a useful marker to identify women who are at higher risk of breast cancer.The


Background and objectives: Inflammatory bowel disease (IBD) is associated with an increased risk of developing colorectal cancer as well as some extra-intestinal tumors, but there are still limited data about the risk of prostate cancer (PC). To analyze if there is an increased risk of PC in patients affected by IBD, we performed a systematic review with meta-analysis. Materials and Methods: A Pubmed search of all studies comparing standardized incidence ratio (SIR) or odds ratio (OR) or relative risks (RR) of PC between IBD and non IBD groups, published until March 2020 was conducted. The study protocol was registered on PROSPERO. Twelve studies, mostly population studies, were included. The quality score of these studies, evaluated by the Newcastle–Ottawa Scale, was 7. The heterogeneity was high among the studies in which ulcerative colitis (UC) was considered separate from Crohn’s disease (CD) and in the studies that considered UC and CD together (“IBD-studies”), while it was low in the studies which considered CD separate from UC. Results: The relative risk of developing PC was 1.71  in IBD, 1.10  in CD, and 1.22  in UC. Conclusions: Patients with IBD appear to have a slightly increased risk of PC compared to the general population. Inflammatory bowel disease (IBD) is characterized by a chronic idiopathic gastrointestinal inflammation due to a dysregulated immune response. Based on clinical, histopathological, and endoscopic features, IBD can be differentiated in Crohn’s Disease (CD) and Ulcerative Colitis (UC) . AlthougChronic inflammation is a well-known risk factor for cancer development, and in the literature, it is widely acknowledged that IBD is associated with a higher risk of intestinal cancers, such as colorectal cancer or small bowel adenocarcinoma ,8. SeverProstate cancer (PC) is the second most frequent malignancy in men worldwide, but only limited and controversial data on the risk of PC in IBD patients are available . Indeed,In the present study, we conducted a systematic review with meta-analysis aimed at analyzing the association between IBD and PC.This systematic review with meta-analysis was conducted according to the PRISMA  guidelines . The stuArticles published in English about risks of PC in patients with IBD were identified through PubMed (“All databases”) searches using the terms “(inflammatory bowel disease [Title] OR Crohn’s Disease [Title] OR Ulcerative Colitis [Title]) AND prostate [Title] AND cancer [Title]”. The final date of the search was March 30, 2020.Reference lists from published articles were also employed, such as citing articles on PubMed Central. Titles of these publications and their abstracts were scanned in order to eliminate duplicates and irrelevant articles. The inclusion criteria were: (1) studies comparing the incidence of PC between a group of IBD patients and those without IBD or subjects belonging to the general population; (2) studies reporting standardized incidence ratio (SIR) or odds ratio (OR) or relative risks (RR) and the corresponding 95% confidence intervals (CI); (3) whatever the sample was. The exclusion criteria were: (1) studies which did not estimate the association between IBD and PC risk; (2) studies which did not report the number of PC in the two cohorts; (3) reviews; (4) case reports; (5) meta-analyses.Two authors (D.G.R. and E.C.) independently reviewed titles and abstracts of references retrieved from the literature search and selected potentially relevant studies. The full-text versions of selected studies were then assessed by the two authors to determine whether the inclusion criteria were satisfied. Differences in opinion were solved by discussion until consensus was reached. If an agreement failed to be reached, a third author (G.M.S.) was consulted.The following information was collected: first author, country, publication year, number of patients included, numbers of cancers observed in the cohort, expected numbers in a matched background population, SIRs or OR or RRs, and 95%CI.Q-test’s p < 0.05, I2 > 50%), a random-effect model was used , otherwise a fixed-effect model was employed.Cochran’s Q and I2 statistics were used to estimate heterogeneity across studies. When assessing heterogeneity .The Newcastle–Ottawa Scale (NOS) was used to evaluate the quality of the included studies. The NOS is based on a “star system” and judges study quality according to three perspectives: selection of the study groups, comparability of the groups, and ascertainment of the outcome of interest. The average of quality score was 7 for all the included studies .To identify publication bias, funnel plots were created and visually evaluated. In the absence of publication bias, the distribution of effect estimates in the funnel plot resembles a symmetrical inverted funnel, whereas heterogeneity, publication bias, or chance lead to an asymmetrical funnel plot .We identified 150 studies satisfying the search terms. After screening titles and abstracts, 12 of these underwent full-text review. During this process, four papers were excluded as one study was a case-report, one study did not report the exact number of patients with PC, and two studies did not have a comparison with a non-IBD population. By reviewing the lists of references from the retrieved articles, four other articles were included. Twelve studies were finally included, 10 of them were cohort studies, while two had a case-control design. Among these studies, six evaluated the risk of PC in both CD and UC, three evaluated only the correlation between UC and PC, one explored only the correlation between CD and PC, and two referred to IBD patients, without differentiating between the two types ,21,22,23The flow diagram about the studies’ identification, screening, eligibility, and the number of included studies is reported in Among the eight studies evaluating PC risk in IBD , three dFive studies, instead, evidenced an opposite scenario, pointing out an increased risk of PC in patients with IBD. Kappelman et al., in a Danish population-based study, reported an SIR for PC of 1.21 (IC 1.08–1.35) . The samp = 0.007). As a high heterogeneity was detected , a random effects model was used . As a low heterogeneity was detected , a fixed effects model was used . High heterogeneity was present ; thus, a random effects model was used , we did not evidence a statistically significant increased risk of PC in both UC  and CD . The difference can be explained by the fact that we included the highest number of studies in our analysis. For instance, five studies included in our meta-analysis were absent from the meta-analyses performed by Chen et al. [Our meta-analysis showed a slightly different result: in fact, while we confirmed an increased risk of PC in IBD patients , or the fact that the IBD patients could have an increased ascertainment of PC because they are more frequently subjected to rectal examinations due to the characteristics of the disease. Furthermore, we did not consider concomitant medications, such as immunosuppressive drugs, that could increase the risk of developing cancer, decrease ability to eradicate infections, and induce direct DNA impairment. Furthermore, there was a high heterogeneity among the studies, probably due to the different age of the participants and follow-up periods. Another possible bias could be the age of the patients of the included studies. Generally, the majority of those affected by PC are older men, whereas UC and CD affect a younger population. Despite that, the included studies matched case and control for age and showed also an SIR (standardized incidence ratio): hence, this possible confounder was eliminated.Importantly, our meta-analysis included the highest number of studies published on this subject; we enrolled not only cohort studies, but even case-control ones, and we did a sub-analysis, excluding studies with lower quality, thus confirming the obtained results.Our study found that men with IBD have an increased risk of PC, but larger studies are needed to evaluate the risk in the IBD subgroups, especially in patients with UC. Further research should aim at evaluating the impact of concomitant medications on the risk of developing PC. These results could be useful in evaluating the possibility of implementing tailored screening programs for PC in patients with IBD.


Polycystic ovarian syndrome (PCOS) is a multi-gene hereditary disorder caused by the interaction of certain gene variation with environmental factors. Previous studies have shown that vascular endothelial growth factor (VEGF) gene polymorphisms are associated with the risk of polycystic ovarian syndrome. However, the results of these studies remain controversial. We performed the present meta-analysis aiming to further investigate the potential relationship between VEGF polymorphisms and susceptibility to PCOS.The following databases were systematically searched: PubMed, EMBASE, Web of Science (WOS), China National Knowledge Infrastructure (CNKI), and Wanfang Databases. The correlation between VEGF polymorphisms and PCOS risk was assessed by calculating pooled odds ratios (ORs) and their 95% confidence intervals (95% CIs). Subgroup analyses stratified by ethnicity and source of control were also conducted. Besides, trial sequential analysis (TSA) was done to verify the reliability of the pooled results.10 relevant case-control studies were incorporated in this meta-analysis, involving 1347 PCOS cases and 1378 controls. The VEGF rs2010963 polymorphism was associated with decreased PCOS risk in the whole population and the Asian populations. The VEGF rs3025039 polymorphism was associated with decreased PCOS susceptibility and the Asian populations, but increased risk of PCOS was observed among the Caucasian populations. In addition, the results of trial sequential analysis (TSA) showed the negative correlation between rs2010963 and PCOS risk, obtained by our meta-analysis, was stable and reliable.Overall, different VEGF gene polymorphisms may exert different effects on PCOS susceptibility. The VEGF rs2010963 polymorphism decreases PCOS susceptibility in both the whole population and the Asian populations, and VEGF rs3025039 polymorphism causes lower PCOS susceptibility in the whole population and the Asian populations but higher in the Caucasian populations. Polycystic ovarian syndrome (PCOS) is one of the most common endocrine diseases with complex etiology and pathogenesis, with a prevalence of up to 10% in reproductive age women . It is gVascular endothelial growth factor (VEGF), a homodimeric glycoprotein hormone first purified by Ferrara et al. from the bovine pituitary follicular stellate cells in vitro, is the most active angiogenic factor ever known, which can maintain the differentiation state of vascular endothelial cells and improve microvascular permeability, and play a powerful regulatory role in the occurrence and development of various diseases , 15. It Recent researches have focused on the correlation between VEGF gene polymorphism and PCOS, with rs2010963 (G > C), rs833061 (T > C), rs3025039 (C > T), rs699947 (A > C), and rs1570360 (A > G) extensively investigated for their potential relationship to PCOS risk . HoweverAfter formulating the search strategy, we systematically searched the databases of PubMed, EMBASE, Web of Science (WOS), Chinese National Knowledge Infrastructure (CNKI) and Wan Fang to collect potentially eligible case-control studies related to VEGF gene polymorphisms and the risk of PCOS. The last retrieval was conducted on October 1, 2019. The search terms were as follows:  and (“PCOS” OR “Polycystic ovary syndrome”) and . To avoid the omission of relevant research, references in the eligible literature after retrieval were also browsed and manually retrieved.The studies included in this meta-analysis must meet the following criteria: (a) Case-control studies covered the relationship between VEGF gene polymorphisms and risk of PCOS. (b) The sample size, genotype distribution, and genotype frequency of the case and control groups were directly provided in the document, or the literature contained available original data that could be converted into OR value and 95%CI for the case and control groups. (c) The diagnosis of PCOS patients in the case group must conform to one of the following three diagnostic criteria: NIH, Rotterdam, and AE-PCOS Society –32 (d). Studies were excluded on the basis of the following criteria: (a) There was no clear definition of polycystic ovary syndrome in the literature, and no clear diagnostic criteria were available for enrolled patients in the case group. (b) The specific genotype data of the case group and the control group were not provided, or the genotype frequency of the case group and the control group could not be calculated by the original data in the literature. (c) Conference papers, case reports, literature reviews or animal researches were ruled out.Two reviewers (Jiahui Zhao and Da Li) read screened independently the eligible relevant literature according to the inclusion and exclusion criteria. When the two researchers disagreed about whether or not a study should be included, they would try to resolve it through discussion. If they could not reach a consensus, the final decision would be made by a third researcher (Lisha Tang). For the included literature after screening, the following data were mainly extracted: first author, publication year, nationality, ethnicity, PCOS diagnostic criteria adopted, inclusion criteria of the control group, source of control, genotyping method, single nucleotide polymorphisms (SNPs) investigated in the study, sample size and genotype distribution of the case group and the control group.The literature quality was evaluated by Newcastle-Ottawa Scale (NOS) . The NOS2 statistic. If I2 was lower than 50%, which indicated that the heterogeneity was low, the fixed-effect model would be applied for meta-analysis. If statistical heterogeneity between studies existed (I2 > 50%), the source of heterogeneity would be further analyzed, and the random effect model would be adopted. In addition to overall analysis, subgroup analyses stratified by ethnicity and source of control were also performed. Since all of the included literature adopted the Rotterdam diagnostic criteria for the enrolled patients of case groups, there was no subgroup analyses based on different diagnostic criteria. If there was only one article in a subgroup, no separate analysis would be performed.Meta-analysis was mainly performed using Revman5.3 software by the Cochrane Collaboration. The strength of association between VEGF polymorphisms and susceptibility of PCOS was evaluated by calculating Odds ratios (OR) and its 95% confidence intervals (CIs) under five gene comparison models. For example, the correlation between rs2010963 (G > C) polymorphism and PCOS risk was measured and assessed under the following models: allele model (C vs. G), dominant model (GC + CC vs. GG), heterozygous model (GC vs. GG), recessive model (CC vs. GC + GG) and homozygous model (CC vs. GG). The heterogeneity between studies was quantitatively determined by I2 test to ascertain whether the control group was consistent with Hardy-Weinberg equilibrium (HWE).The possible publication bias was first evaluated by observing the symmetry of the funnel plots. The Stata14.0 software  was further applied to perform the Begg’s and Egger’s test procedures to assess potential publication bias. The genotype distribution of the control group was calculated by χTraditional meta-analysis methods have no way to assess the adequacy of the overall sample size of the included studies. When the overall sample size is not sufficient, the final pooled results obtained are not reliable enough, especially the positive results obtained by meta-analysis. Statistical results based on insufficient sample size increase the likelihood of false positive results (Type 1 error). Therefore, in the present study, we adopted trial sequential analysis (TSA) method to determine whether the overall sample size of the included studies was sufficient and whether the overall results obtained were reliable. Trial sequential analysis (TSA) was performed by using the TSA v0.9.5.10 Beta software . For theThe meta-analysis was performed by summarizing data from previous studies. Therefore, no ethical approval was required.A total of 129 articles were obtained through initial literature retrieval. After excluding 25 duplicates, the remaining 104 papers were filtrated by reading their titles and abstracts. After further screening, totally 12 papers entered the full-text acquisition and detailed screening. After excluding a repetitive study based on the same case group and another article focusing on the association between VEGF gene polymorphisms and insulin resistance in patients with PCOS, 10 articles met the inclusion criteria and eventually merged into the study. A total of 1347 PCOS cases and 1378 controls were included in the 10 included studies –29. Thisp value of Hardy-Weinberg equilibrium (HWE) in each control group. The Newcastle-Ottawa Scale scores for all the included studies are shown in Table The flowchart of selection and inclusion process is displayed in Fig. P = 0.008, I2 = 49%; Heterozygote model: OR = 0.75, 95%CI: 0.61–0.91, P = 0.04, I2 = 48%). When the studies were stratified by ethnicity, significantly decreased risk of PCOS was observed in the Asian populations under 3 gene comparison models . No significant association was found between VEGF rs2010963 polymorphism and PCOS susceptibility in the Caucasian ethnicities and stratified subgroup analyses based on source of control. Figure The main results of the overall and subgroup meta-analysis for VEGF gene polymorphisms and PCOS susceptibility are presented in Table Six studies had 936 cases and 1010 controls for VEGF rs2010963 polymorphism, 3 of which were based on the Asian populations and another 3 on the Caucasian populations. There was no significant correlation found between VEGF gene rs833061 polymorphism and risk of PCOS, whether based on aggregate analysis of all populations or subgroup analyses based on different ethnicities and sources of control.P = 0.05, I2 = 0%), Based on ethnic stratification, significantly increased risk of PCOS was observed in the Caucasian populations under 2 gene comparison models . However, in the Asian populations, there was negative correlation between VEGF rs2010963 polymorphism and PCOS risk under Homozygote model . Moreover, subgroup analysis based on source of control indicated that VEGF rs3025039 polymorphism was related to decreased PCOS risk in the population-based group under Homozygote model .Five studies had 769 cases and 779 controls for VEGF rs3025039 polymorphism, 3 of which were based on the Asian populations and 2 on the Caucasian populations. Overall, the pooled results based on the whole population revealed significantly negative correlation between VEGF rs2010963 polymorphism and PCOS risk under Homozygote model , suggesting that the accumulated sample size was sufficient and the pooled positive result was credible. Under the heterozygote model of rs2010963, the cumulative Z curve passed through the traditional boundary threshold, meanwhile crossed the trial sequential monitoring boundary, and confirmed a reliable conclusion although the cumulative sample size did not transcend the required information size (RIS). In other gene comparison models that yielded positive results, the cumulative Z curve did not cross the sequential trial monitoring boundary, and the sample size did not reach RIS, demonstrating that the pooled results possibly were not solid enough. Figure Polycystic ovary syndrome (PCOS) is a female endocrine disorder with complex etiology. Currently, the widely adopted diagnostic criteria are Rotterdam criteria: rare ovulation or anovulation, clinical or biochemical hyperandrogenemia, and ovarian polycystic changes determined by ultrasound. Patients with two of three in the criteria and without other diseases that may cause hyperandrogenemia could be diagnosed with PCOS . The proVEGF protein is known to be closely related to the pathogenesis of PCOS. Serum VEGF levels are significantly elevated in PCOS patients –18. HighIt must be recognized that this meta-analysis has some possible limitations. First, the number of articles included in the study was small, and the overall number of patients in the case group and control group was low. Some subgroup analyses were unable to perform data analysis because only one study was included, thus limiting the reliability of summary results. Secondly, the previous studies on VEGF gene polymorphisms and PCOS risk only contained two ethnicities, lacking relevant studies on other races. Third, although the inclusion criteria for the control group were clearly stated in all studies, the inclusion criteria for different studies were not the same. For example, in some studies, the inclusion criteria for the control group were healthy women of childbearing age, while in others the criteria detailed a history of successful childbearing and a recent absence of any medications that might affect hormone levels. As PCOS is a disease with complex etiology, its incidence is related to many factors, such as obesity and insulin resistance. Due to the absence of relevant patient data based on the two factors in the original study literature, we did not conduct subgroup stratification analysis based on factors such as whether the patient was obese or not and the severity of insulin resistance.Despite some limitations, this meta-analysis still has several highlights. To date, this is the first known meta-analysis focused on the correlation between VEGF gene polymorphisms and PCOS risk. Through the establishment of strict inclusion and exclusion criteria, a total of 10 studies were included in this meta-analysis, involving 1347 PCOS cases and 1378 controls. The study spanned seven regions and included both Caucasian and Asian populations. We systematically evaluated the effect of VEGF gene polymorphism on the occurrence of PCOS. At the same time, in most gene comparison models, the calculated inter-study heterogeneity was low, indicating the results of this meta-analysis were stable. In addition, the results of trial sequential analysis (TSA) also demonstrate the correlation between rs2010963 and decreased risk of PCOS obtained through this current meta-analysis was stable and reliable. Many previous studies have suggested that multiple SNPs of VEGF gene increase the risk of diseases such as ovarian malignancies, hypertensive disorder of pregnancy, and idiopathic recurrent pregnancy loss, and thereafter have been confirmed by multiple meta-analyses. Our meta-analysis results revealed certain VEGF gene polymorphisms had a distinct different impact on the pathogenesis of PCOS in contrast to its role in the pathogenesis of other female reproductive system disorders –45. TherIn conclusion, this study establishes a significant correlation between the variation of VEGF rs2010963 and decreased PCOS susceptibility in the entire population; subgroup analysis shows this association only occurs occurred in Asian populations. The mutation of VEGF gene rs3025039 may also decrease the risk of PCOS in the overall population; subgroup analysis based on different populations shows that rs3025039 polymorphism is also associated with decreased risk of PCOS in the Asian populations, while in the Caucasian populations rs3025039 polymorphism of VEGF gene enhances the risk of PCOS. More well-designed studies involving more cases will be needed in the future to confirm this inference.


Observational studies have suggested that polyunsaturated fatty acids (PUFAs) may decrease Alzheimer’s disease (AD) risk. In the present study, we examined this hypothesis using a Mendelian randomization analysis.We used summary statistics data for single-nucleotide polymorphisms associated with plasma levels of n-6 PUFAs  and n-3 PUFAs , and the corresponding data for AD from a genome-wide association meta-analysis of 63,926 individuals .None of the genetically predicted PUFAs was significantly associated with AD risk; odds ratios  per 1 SD increase in PUFA levels were 0.98  for linoleic acid, 1.01  for arachidonic acid, 0.96  for alpha-linolenic acid, 1.03  for eicosapentaenoic acid, 1.03  for docosapentaenoic acid, and 1.01  for docosahexaenoic acid.This study did not support the hypothesis that PUFAs decrease AD risk.The online version of this article (10.1007/s00394-019-02126-x) contains supplementary material, which is available to authorized users. The overall results from observational studies suggest that daily intake of polyunsaturated fatty acids (PUFAs) is inversely associated with risk of incident Alzheimer’s disease (AD) and cognitive impairment . PreviouHowever, observational studies are generally susceptible to methodological problems such as confounding. Therefore, randomized controlled trials (RCTs) are more suitable to examine causal relationships. However, no RCT has yet reported the preventive effect of PUFAs on incident AD among cognitively normal adults .To overcome the problem of confounding in observational studies, Mendelian randomization (MR) is becoming widespread for assessing causal relationships. MR study is a type of instrumental variable analysis where genetic variants (single-nucleotide polymorphisms: SNPs) are applied as instrumental variables for the potential risk factor. MR study is often described as a “natural RCT” because random allocation of alleles during meiosis is conceptually similar to the RCT design. Therefore, a MR study would provide more robust evidence regarding the causal relationship between PUFAs and AD than observational studies. However, to our knowledge, no MR study has yet investigated this relationship.The aim of the present MR study was to examine the hypothesis that higher plasma levels of PUFAs decrease AD risk. We examined the main six types of PUFAs, including the n-6 PUFAs linoleic acid (LA) and arachidonic acid (AA) and the n-3 PUFAs alpha-linolenic acid (ALA), eicosapentaenoic acid (EPA), docosapentaenoic acid (DPA), and docosahexaenoic acid (DHA).To estimate causal associations between lifelong exposures to PUFA levels and the risk of late-onset AD, we conducted a two-sample MR analysis using summary statistics data from genome-wide association studies (GWAS).n = 8631 individuals for n-6 PUFAs and 8866 individuals for n-3 PUFAs). Age ranged from 21 to 102 years [We used summary statistics data from the Cohorts for Heart and Aging Research in Genomic Epidemiology (CHARGE) consortium examining the association between genetic variants and plasma levels of n-6 PUFAs  and n-3 We used two types of summary statistics data for AD.First, to conduct primary analysis, we used data from GWAS for the association between genetic variants and clinically diagnosed late-onset AD (Table S1) . The sumrg = 0.81) [Second, to conduct secondary analysis, we used summary statistics data from another GWAS for the association between genetic variants and AD including AD-by-proxy based on parental diagnoses (Table S2) . The sumP < 5 × 10−8); (2) biologically relevant to PUFAs (involved in PUFA metabolism); (3) strongest association within the specific locus  [F statistic ≥ 10) [We selected 8 SNPs associated with one or more of the PUFAs (Table S1). All SNPs satisfied the following conditions: (1) associated with the PUFAs at genome-wide significance P <  × 10−8; g FADS1) . In addiic ≥ 10) .For each PUFA, two-sample MR analyses were performed to calculate the odds ratios and 95% confidence intervals for AD. Odds ratios were shown as per one standard deviation increase based on standard deviations of the largest study (the Atherosclerosis Risk in Communities Study) in the CHARGE consortium (Table S3) , 10. AllBecause the present study is based on GWAS summary statistics rather than individual level data, an ethical approval is not required according to the rules at Karolinska Institutet, Sweden.The results of the MR analyses based on the inverse variance weighted method are shown in Fig. P-intercept > 0.05).Results obtained by the weighted median and MR-Egger regression analyses (Table S4) were not essentially different from those based on the inverse variance weighted method Fig. . Plots tResults from the secondary analyses (Table S5) were consistent with the main findings Fig. .To our knowledge, this is the first MR study to examine the relationship between genetically predicted levels of PUFAs and AD. Our results showed that none of the PUFAs was statistically significantly associated with AD risk. Results from our MR study, which is less prone to confounding compared with observational studies, did not support previous results from observational studies showing that PUFAs are inversely associated with AD risk.F statistic > 10) [As major methodological issues about bias towards the null in MR studies, weak instrument bias and low statistical power are often suggested. Although we only used 1–3 SNPs as instrumental variables for each PUFA, the SNPs explained a relatively large variation in PUFA levels and they fulfilled the criterion as not being weak instrumental variables (ic > 10) . AdditioP < 0.05) [P < 0.001) [Regardless of statistical power, effect sizes (point estimates) were smaller in the current MR study than in previous observational studies. For example, a recent cohort study reported that the hazard ratio per 1 SD increment of red blood cell EPA + DHA levels for dementia was 0.92 ( < 0.05) . FurtherThe present study has several limitations. First, we only used 1–3 SNPs as instrumental variables for each PUFA. Therefore, regression estimates such as pleiotropy test by MR-Egger regression might not be robust. Second, because shared SNPs were used among the PUFAs , identifying the effects for individual PUFAs was difficult.In conclusion, the present MR study did not support the hypothesis that PUFAs decrease AD risk.Supplementary material 1 (DOCX 370 kb)Below is the link to the electronic supplementary material.


Background: Periodontitis and metabolic syndrome (MetS) are two major global health problems that are widely prevalent in the world, although the former is a common infection in developing countries and the latter is a non-infectious but prevalent disease in developed countries. This study aims to provide an updated review on the existence and magnitude of the relationship between periodontal disease and the risk of MetS.Methods: We searched the PubMed, Web of Science, ScienceDirect, Chinese National Knowledge Infrastructure, and Wanfang databases for original studies assessing the association between periodontitis and MetS published before August 2019. We calculated the pooled crude and adjusted odds ratios (ORs) together with the 95% confidence intervals (95% CIs) to estimate the strength of this association. Subgroup analysis was performed by considering the diagnostic method or the country where the studies were performed.Results: We identified 43 potentially eligible articles for this systematic review, including 32 cross-sectional studies, eight case–control studies, and three cohort studies. Among them, 39 articles presented enough information to be included in the meta-analysis. The pooled crude and adjusted ORs were 1.99 (95% CI: 1.75–2.25) and 1.46 (95% CI: 1.31–1.61), respectively. Subgroup analysis showed a consistent relation stratified by either the diagnostic method or the country where the studies were performed. The pooled OR was 1.68 (95% CI: 1.41–2.00) for Japan, 1.75 (95% CI: 1.31–2.34) for the USA, 1.81 (95% CI: 1.35–2.42) for Korea, and 2.29 (95% CI: 1.53–3.41) for China.Conclusion: Our results provide compelling evidence for the association between periodontitis and MetS. Patients with periodontal disease are a critical screening population for MetS. We also recommend that people exhibiting components of MetS should receive a periodontal check-up and pay attention to their oral health. Oral health is an undervalued parameter of global health and has been considered inferior on the agendas of policymakers . There hMetS is a prevalent and multifactorial disorder that consists of a cluster of several clinical physical conditions and biological abnormalities that increase the risk of mortality; MetS is affected by insulin resistance and promotes cardiovascular diseases . These cThe association between periodontitis and MetS has gained research interest in the scientific literature. There is variation in the reported degree of association, which may be due to the different definitions of MetS, the methodology used to assess periodontitis, and criteria for subject enrollment . In addiThe PRISMA  guidelines were used  13). We. We13). Two reviewers independently identified eligible studies on the above topic, assessed the publication validity, and subsequently extracted data. The discrepancies were discussed and resolved, and the studies were concluded after consultation with the supervisor of this research. We collected data on the year of publication, the first author, study country, study design, sample size, mean age, sex, current tobacco smoking status, criteria for periodontitis diagnosis, periodontal examination protocol, the number of subjects diagnosed with periodontitis, criteria for MetS diagnosis, the number of subjects diagnosed with MetS, and the strength of the association .We used the Newcastle–Ottawa scale to evaluate the quality of the studies. A “star system” was used to judge the study in the following three contexts: the selection of the study groups, the comparability of the groups, and the ascertainment of either the exposure or the outcome of interest. The results varied across the selected studies, which is shown in I2 was also used to test the heterogeneity among the included studies. Consequently, subgroup analysis was first performed on the method used to diagnose periodontitis, that is, either partial-mouth periodontal examination or full-mouth periodontal examination. The second subgroup analysis was on the criteria for the diagnosis of MetS, which are the National Cholesterol Education Program Adult Treatment Panel III (NCEP ATP III) criteria, the 2005 International Diabetes Federation  criteria, or the 2009 IDF criteria. Finally, the subgroup analysis was performed by country, namely, China, Japan, the USA, and Korea, the countries where most of the published studies have been conducted. The subgroup analysis was performed using both the crude and the adjusted ORs separately. Stata version 15.0  was used to analyze the data.The associations between periodontitis and MetS were assessed by the pooled ORs with the corresponding 95% CIs. Data from 39 studies were collected individually, and a crude OR was calculated, followed by the random-effect model. Then, the second analysis, consisting of the pooled adjusted OR, was derived from the given adjusted ORs from each study. During the literature search, 1,125 articles were selected, and to find other associated studies, articles listed in their references were also screened. Duplicates and other non-related articles were not considered for the systematic review. Then, among these, 86 of the articles were retrieved, and their titles, abstracts, and full texts were scrutinized for possible relevance. Finally, 43 articles were selected for the systematic review, and 39 were enrolled in this meta-analysis where the crude ORs were calculated individually . AdditioThe characteristics of each study are listed in I2 = 87.7% and P < 0.001 . The heterogeneity test showed that  < 0.001 . We furt < 0.001 .I2 = 94.0%, P < 0.001) for partial periodontal examination and a crude OR of 2.11  for complete periodontal examination  for the complete periodontal examination and 1.47  for the partial periodontal examination  for the 2009 IDF criteria, 2.08  for the NCEP ATP III criteria, and 2.18  for 2005 IDF criteria  for the NCEP ATP III criteria, 1.48  for the 2009 IDF criteria, and 2.39  for the 2005 IDF criteria  for Japan, 1.75  for the USA, 1.81  for Korea, and 2.29  for China  for the USA, 1.41  for Korea, 1.51  for China, and 1.81  for Japan, respectively  and 4.2 (95% CI: 1.2–14.8), respectively (Our results provide compelling evidence for the association between periodontitis and MetS. Patients with periodontal disease are a critical screening population for MetS. We also recommend that people exhibiting components of MetS should receive a periodontal check-up and pay attention to their oral health. Among the golden rules for maintaining good oral hygiene is that one should visit a dentist at least twice a year, brush at least twice daily, floss regularly, and have a healthy diet, thereby avoiding too many sweet and sticky foodstuffs. Particular attention should be given to children, and oral hygiene, as well as healthy lifestyle habits, should be implemented in the educational curriculum so that from an early age, one is made aware of these basics to have a healthier and brighter future generation. Policymakers should invest in the promotion and prevention of these issues so that the financial impact due to the treatment and rehabilitation of these diseases is less.RG, DT, and JW conceived the study, analyzed the data, and drafted the manuscript. QL participated in the study design and helped refine the manuscript. All authors contributed to the study and have read and approved the final version.The authors declare that the research was conducted in the absence of any commercial or financial relationships that could be construed as a potential conflict of interest.


Long noncoding RNAs (lncRNAs) have been widely studied, and single nucleotide polymorphisms (SNPs) in lncRNAs are considered to be genetic factors that influence cancer susceptibility. The lncRNA GAS5, MEG3, and PCAT-1 polymorphisms are shown to be possibly associated with cancer risk. The aim of this meta-analysis was to systematically evaluate this association.  Studies were selected from PubMed, Web of Science, Embase, Google Scholar, Cochrane Library, the Chinese National Knowledge Infrastructure (CNKI), and the Chinese Biomedical Literature Database (CBM) through inclusion and exclusion criteria. Odds ratios (ORs) and 95% confidence intervals (CIs) were calculated using the random-effects model or fixed-effects model to assess the association between lncRNA polymorphisms and cancer susceptibility. Metaregression and publication bias analyses were also conducted. All analyses were performed using the Stata 12.0 software. del vs. ins: OR = 0.79, 95%CI = 0.72‐0.86; del/del vs. ins/ins+del/ins: OR = 0.74, 95%CI = 0.59‐0.91; del/ins vs. ins/ins: OR = 0.84, 95%CI = 0.67‐1.05). For rs16901904, a decreased cancer risk was observed in three genetic models . No statistical significance was found in the metaregression analysis. For all of the included SNPs, no publication bias was found in all genotype models.  Sixteen articles (covering 13750 cases and 17194 controls) were included in this meta-analysis. A significant association between SNP rs145204276 and gastric cancer risk was observed ( The rs145204276 SNP in lncRNA GAS5 is likely to be associated with gastric cancer risk, whereas the rs16901904 SNP in lncRNA PCAT-1 bears association with a decreased cancer risk. Cancer has become a leading cause of death worldwide not only in high-income but also in middle-income countries , 2. Due High-throughput sequencing technologies have identified a great number of noncoding RNAs in the genome, which could be classified into small 18-200 nts) and long (200 nts to >100 kb) noncoding RNAs (lncRNAs) . Althoug-200 nts Recently, research showed that GAS5 overexpression could inhibit the invasive ability of hepatocarcinoma cells affecting the epithelial-mesenchymal transition (EMT) process, which is very important in early events of the metastatic spread of tumor cells and can make cells more active and invasive . MeanwhiFurthermore, lncRNAs such as GAS5, H19, MEG3, and TUSC7 play oncogenic or tumor suppressor roles in correlation with tumor suppressor protein P53 or oncoprotein c-Myc, respectively . Balci eTwo investigators independently conducted an electronic literature search  using PubMed, Web of Science, Embase, Google Scholar, Cochrane Library, the Chinese National Knowledge Infrastructure (CNKI), and the Chinese Biomedical Literature Database (CBM). The restriction in publication languages was selected as English and Chinese. During our search process, the following keywords were used: “lncRNA GAS5,” “lncRNA MEG3,” “lncRNA PCAT-1,” “polymorphism,” “Single nucleotide polymorphism,” “genetic variant,” “cancer,” “tumor,” “malignancy,” “carcinoma,” and “neoplasm.” All clearly irrelevant studies such as case reports and review articles were excluded. The relevant studies cited in the references of review articles were also manually searched for additional eligible studies.Two researchers independently evaluated the titles and abstracts of the identified articles to decide whether they met the study criteria. Differences were solved by consensus. Both cross-sectional and longitudinal studies on the relationship between SNPs in lncRNAs GAS5, MEG3, and PCAT-1 and cancer risk were included. Studies were included in this meta-analysis only when they met the following criteria: (1) the study evaluated the association of SNPs in lncRNAs GAS5, MEG3, and PCAT-1 with cancer risk; (2) the study was published in Chinese or English language; (3) the study had an original case-control design or was a cohort-designed study in humans; (4) the study provided sufficient genotyping data to estimate ORs and 95% CIs; (5) genotype frequencies of subjects in controls were in accordance with the Hardy-Weinberg equilibrium (HWE); (6) all cancer cases were diagnosed at any stage by pathology, but only primary tumor could be involved. If there was a parallel publication, we selected the study with a larger sample size.Studies were excluded if (1) they were not case-control studies; (2) they were case reports, comments, meta-analyses, or review articles; (3) they were a previous study that was duplicated; (4) they were the control groups that did not conform to HWE; (5) there were not enough genotyping data to calculate the ORs and 95% CIs; and (6) they were studies on cell lines and gene expression. If there was only one study investigating one particular lncRNA gene which was not suitable for meta-analysis, it was also excluded. Additionally, when researchers did not report detailed information regarding the genotype distribution in each group, the corresponding authors of the study were contacted for unpublished data.P value of HWE in controls. Any inconsistencies were resolved by a third researcher until a consensus was reached.Data extraction was performed independently by two researchers, including the following information from each study: first author, year of publication, ethnicity, cancer type, source of controls, number of cases and controls, genotype distribution of cases and controls, genotyping method, and The present meta-analysis investigates the relationship between lncRNA GAS5, lncRNA MEG3, and lncRNA PCAT-1 SNPs and various types of cancers with ORs and 95% CIs. For each SNP, five genetic models of ORs and 95% CIs  were estimated. Subgroup analysis was conducted according to ethnicity, source of controls, and cancer types.P > 0.05) was applied to calculate the P value of HWE in control groups. The associations between each SNP and cancer susceptibility were estimated by pooled odds ratios (ORs) and 95% confidence intervals (CIs) under five different genetic models. The heterogeneity of results throughout the studies was assessed using the I2 statistic, which describes the percentage of total variation across the studies that was attributable to heterogeneity rather than to chance. The I2 values of 25% and 75% corresponded to cutoff points for low and high degrees of heterogeneity. The pooled effect was calculated using the random-effects model when the I2 value was >75%. Otherwise, a fixed-effects model was used in case of significant heterogeneity across studies. A sensitivity analysis was performed to evaluate the influence of each individual study on overall estimates. Begg and Egger's tests were used to assess potential publication bias. All statistical analyses were conducted by Stata software version 12.0 . A P value < 0.05 was considered statistically significant.Chi-square test  were included in the quantitative synthesis. A flowchart of the study selection process is shown in The characteristics of the studies included in this meta-analysis are presented in del vs. ins: OR = 0.79, 95%CI = 0.72‐0.86; del/del vs. ins/ins+del/ins: OR = 0.74, 95%CI = 0.59‐0.91; del/ins vs. ins/ins: OR = 0.84, 95%CI = 0.77‐0.98) , as shown in The relationship between the rs145204276 SNP in the lncRNA GAS5 gene and the risk for all types of cancers is shown in 77‐0.98) . A simil77‐0.98) . AlthougAA vs. GG: OR = 0.86, 95%CI 0.46‐1.63; AG vs. GG: OR = 0.98, 95%CI = 0.73‐1.33; AA+AG vs. GG: OR = 0.97, 95%CI = 0.69‐1.36; AA vs. AG+GG: OR = 0.87, 95%CI = 0.51‐1.47)  .C vs. T: OR = 0.79, 95%CI = 0.70‐0.90; CC vs. CT+TT: OR = 0.49, 95%CI = 0.37‐0.65; CC vs. TT: OR = 0.49, 95%CI = 0.37‐0.66). Similar associations were found in rs710886, although not statistically significant , shown as For lncRNA PCAT-1 polymorphisms, we mainly focused on investigating the effects of four lncRNA polymorphisms . As shown in I2 value > 75%) was observed in the overall or subgroup analyses, the random-effects model was used to make stable confidence intervals. Otherwise, the fixed-effects model was selected to conduct data analysis.We observed a significant heterogeneity among studies in the meta-analysis on the rs145204276 polymorphism. We hence performed a metaregression analysis to assess the source of heterogeneity in the del vs. ins genetic model . The canP value of Begg's funnel plot test was 0.78.Begg's funnel plot and Egger's test were performed to examine publication bias in this meta-analysis. For all included SNPs, no evidence of publication bias was found in all genotype models. One of the publication bias test results is shown as In recent years, lncRNAs have been confirmed to play important regulatory roles in gene expression, and their aberrant expression has been recognized as a hallmark feature of cancer . SNPs hadel vs. ins: OR = 0.79, 95%CI = 0.72‐0.86; del/del vs. ins/ins+del/ins: OR = 0.74, 95%CI = 0.59‐0.91; del/ins vs. ins/ins: OR = 0.84, 95%CI = 0.77‐0.98). Nevertheless, we did not detect an association of the rs145204276 polymorphism with risks for other cancers. Interestingly, the mechanism of action of the rs145204276 polymorphism in cancer cells has also not yet been completely elucidated. On the one hand, researchers observed that the del polymorphism rs145204276 may influence GAS5 transcriptional activity by affecting the methylation status of a CpG island in the promoter region of GAS5, which affects its tumor-suppressing function [loci are different in different cancer types. The mechanism of action of GAS5 in cancer is still poorly understood, which also requires further studies.Recently, lncRNA GAS5 has attracted attention as a new type of lncRNA which plays a key role in cancer development. Moreover, the 5 bp indel polymorphism (rs145204276) in the lncRNA GAS5 promoter region has also been studied in detail in different types of tumors. However, we could not obtain an association between the rs145204276 polymorphism and cancer risk in none of the five genetic models in the overall analysis. The results of the subgroup analysis by cancer type revealed that the rs145204276 del allele was associated with a decreased risk of gastric cancer . Similar associations were observed in rs710886, although not statistically significant. This finding still needs to be further studied.lncRNA PCAT-1 is a long intergenic noncoding RNA transcript from the 8q24.21 region . It has Although significant associations between rs145204276 and rs16901904 and cancer risk were observed in this meta-analysis, our current work still presents several limitations. Firstly, there is a high degree of heterogeneity among the included studies on rs145204276. Although we stratified our analysis by cancer type, ethnicity, and source of controls, between-study heterogeneity was still obvious. Metaregression also revealed that the source of heterogeneity was not the ethnicity, genotyping method, cancer type, or source of controls. Secondly, due to lack of original data of the included studies, the evaluation of associations between age, gender, family history, or environmental factors and cancer risk was not conducted. Thirdly, there was only one study in Caucasian population, and the results of this meta-analysis could only be suitable for the Chinese population. Finally, some sample sizes of the included studies were small, which may influence the results' stability.Our results indicate that rs145204276 in lncRNA GAS5 is a protective factor in the development of gastric cancer, and the role of rs145204276 was also distinct in different cancer stages. In addition, rs16901904 in lncRNA PCAT-1 also plays a protective role against cancer development in Chinese populations. However, several limitations still exist in our analysis, so results should be regarded with caution. Larger and multiple ethnicity studies should be included in our further studies.


Conclusion: The findings confirmed that the NSCL/P risk was significantly associated with the G allele and GG genotype of rs560426 polymorphism but not for rs481931 polymorphism. There were no associations between both polymorphisms (rs560426 and rs481931) and the NSCL/P risk in those of European descent and the mixed ethnicities.Background: A number of genes are associated with the incidence of non-syndromic cleft lip/palate (NSCL/P). Studies have shown a significant association between polymorphisms of ATP-binding cassette, sub-family A, member 4 (ABCA4) with the risk of NSCL/P. The present meta-analysis assessed the association between ABCA4 polymorphisms (rs560426 and rs481931) and the NSCL/P risk by reviewing case-control studies. Methods: Four databases  were searched for articles published up to June 2020. The Review Manager 5.3 software was used to calculate the crude odds ratio (OR) and 95% confidence interval (CI). Both subgroup analyses for ethnicity and source of controls and a meta-regression related to publication year were conducted. Results: Of 94 retrieved studies, 12 were analyzed in this meta-analysis (2859 NSCL/P patients and 3792 controls for ABCA4 rs560426 polymorphism and 1333 NSCL/P patients and 1884 controls for ABCA4 rs481931 polymorphism). Overall, there was no significant association between both polymorphisms and the risk of NSCL/P. However, subgroup analysis demonstrated that there was a higher risk of NSCL/P for specific models: the allelic model (OR = 1.13;  More than 70% of all cleft lips/palates (CL/Ps) present without further anomalies and are thus nonsyndromic . Its pre−8 and 5.01 × 10−12 for the rs560426 polymorphism in combined European and Asian cases and another study [−12 for the rs481931 polymorphism in Asian families. A third study [−12 for rs560426 in an Asian population. In addition, some studies showed a significant association between rs560426 and rs481931 polymorphisms of ABCA4 with the risk of NSCL/P [ATP-binding cassette, sub-family A, member 4 (ABCA4) belongs to the transmembrane protein superfamily , which ier study  reportedrd study , a genomf NSCL/P ; in contf NSCL/P . Given tf NSCL/P . There aThis report conforms to the Preferred Reporting Items for Systematic reviews and Meta-Analyses (PRISMA) guidelines .ABCA4” or “ATP-binding cassette subfamily A member 4” or “rs560426” or “rs481931”) and . In addition, we checked the references of retrieved articles, review articles, and GWAS studies to find potential articles.Four electronic databases  were searched for articles published until 21 June 2020, without restrictions of language or publication date. The searched terms were: (“ABCA4 rs560426 (A > G) and/or rs481931 (C > A) polymorphisms, and (IV) having the required data to calculate the odds ratios (ORs) with 95% confidence intervals (CIs). The studies were excluded if they did not have the required data regarding genotype distributions or were animal studies, review articles, letters to the editor reporting previous studies, and family-based studies. The second reviewer (M.M.I.) checked the relevant articles based on the eligibility criteria. The differences between the two reviewers were resolved by the third reviewer (S.K.T.).One reviewer (M.S.) retrieved the studies from the databases, excluded the duplicate and irrelevant studies after reviewing the titles and abstracts. This was followed by full-text review of the eligible articles. The studies were included if they met the following inclusion criteria: (I) case-control design; (II) NSCL/P was the outcome of interest; (III) reporting p-value of the Hardy–Weinberg equilibrium in controls. If there was a disagreement between the reviewers, the problem was solved by a third reviewer (S.B.).Two reviewers (M.M.I. and M.S.) independently extracted the data from each study. The information retrieved from the studies included: the first author’s name, the publication year, the ethnic group, the source of controls, the mean age, the number of males in the two groups, the number of patients and controls with each genotype, the genotyping method, and the One reviewer (M.M.I.) rated the quality of each included article using the Newcastle–Ottawa Quality Assessment Scale (NOS). The NOS involves assessment of each article for quality in three components: selection, comparability, and exposure. A case-control study could have a minimum and maximum total score of 0 and 9, respectively; a higher score demonstrates better quality .ABCA4 polymorphisms and the risk of NSCL/P in the five genetic models [2 statistic. Heterogeneity was considered to be statistically significant if p < 0.1 or I2 > 50%. If there was no significant heterogeneity, the fixed-effects model (Mantel–Haenszel method) was used to estimate the values. Otherwise, we used the random-effects model (DerSimonian and Laird method). The Chi-square test was used to calculate the Hardy–Weinberg equilibrium in the control group of each study. Subgroup analysis was done according to the ethnicity and the source of controls. Meta-regression is a quantitative method used in meta-analysis to estimate the impact of moderator variables on the study effect size. The Comprehensive Meta-Analysis 2.0 was used to derive a funnel plot using the Egger’s and Begg’s tests and p < 0.05 indicated significant existence of publication bias. To evaluate the stability of the results, the following sensitivity analyses were applied: “cumulative analysis” and “one study removed”. The analysis was checked by three reviewers in a discussion .An analysis was performed by the Review Manager 5.3 using crude OR and 95% CI to show the association between c models . The Z tABCA4 polymorphisms and NSCL/P risk or irrelevant studies (11 studies), a total of 23 studies were evaluated  and excluding the book chapters, conference papers, and studies without reporting the association between the valuated . Of 23 sABCA4 polymorphisms in the Asian ethnicity, four [ABCA4 rs560426 polymorphism, but only six [ABCA4 rs481931 polymorphism. There were different genotyping methods among the studies presented in ABCA4 rs560426 polymorphism and two studies [ABCA4 rs481931 polymorphism. We included these studies in the analyses because by excluding them, the number of studies would have been reduced, while this could have been one important reason for an overall biased pattern of results. There were 2859 NSCL/P cases and 3792 controls in the meta-analysis of the ABCA4 rs560426 polymorphism and 1333 NSCL/P cases and 1884 controls in the meta-analysis of the ABCA4 rs481931 polymorphism.ty, four ,9,23,31 ty, four ,33,34 inty, four ,30,32,33ty, four ,31,34, tAll studies were scored for quality , and allABCA4 rs560426 polymorphism from the twelve studies. The heterogeneity in allelic, homozygote, heterozygote, and dominant models was high (I2 > 50%), and therefore the random-effects model was used. In the recessive model, the fixed-effects model was used because of low heterogeneity (I2 < 50%). The pooled OR was 1.01 ) in the allelic model (G vs. A), 1.08 ) in the homozygote model (GG vs. AA), 0.93 ) in the heterozygote model (AG vs. AA), 0.89 ) in the dominant model (AG + GG vs. AA), and 1.08 ) in the recessive model (GG vs. AA + AG). Overall, individuals with the ABCA4 rs560426 polymorphism were not at higher risk for NSCL/P than those without the ABCA4 rs560426 polymorphism.ABCA4 rs481931 polymorphism based on the findings from the six studies are shown in 2 < 50%). The pooled OR was 0.92 ) in the allelic model (A vs. C), 0.85 ) in the homozygote model (AA vs. CC), 0.88 ) in the heterozygote model (CA vs. CC), 0.87 ) in the dominant model (CA + AA vs. CC), and 0.89 ) in the recessive model (AA vs. CC + CA). Based on the genetic models, there was no significant risk of NSCL/P in those with the ABCA4 rs481931 polymorphism.The pooled results for the risk of NSCL/P related to the ABCA4 rs560426 polymorphism (p = 0.03), the homozygote model , and the recessive model . The results did not demonstrate a risk of NSCL/P related to the rs560426 polymorphism among people of European descent and mixed ethnicities (such as Brazilian and Mexican ethnicities). Furthermore, the source of controls did not seem to have any effect on the association of NSCL/P and the ABCA4 rs560426 polymorphism.The first subgroup analysis evaluated the effect of ethnicity and the source of controls in the association of NSCL/P with the morphism . A signiABCA4 rs481931 polymorphism according to the ethnicity and the source of controls . The quality score of all studies was high (≥7) and therefore we could not use a sensitivity analysis  for this topic.The sensitivity analyses , identified that the pooled ORs under all genetic models were stable and trustworthy . Further, excluding two studies ,28 with  studies  did not ABCA4 polymorphisms and the risk of NSCL/P using the five genetic models. The Egger’s and Begg’s tests did not reveal any publication bias (p > 0.05) except for four states. Both tests for allele and homozygote models revealed a publication bias and Egger’s test for heterozygote and recessive models revealed a publication bias (p < 0.05).ABCA4 and the risk of NSCL/P based on five genetic models. The key findings were as follows: While there was no significant association between both polymorphisms and the risk of NSCL/P, in contrast, subgroup analyses demonstrated that there was a higher risk of NSCL/P for the allelic model, the homozygote model, and the recessive model in the Asian ethnicity for the rs560426 polymorphism. Further, there were no associations between both polymorphisms (rs560426 and rs481931) and the NSCL/P risk in the European descent and the mixed ethnicities. The present results showed association of the rs560426 polymorphism and the risk of NSCL/P among the Asian, but not among the European, descent. Further, the meta-regression showed that the number of participants was a confounding factor for the association between the rs481931 polymorphism and the risk of NSCL/P. Overall, the ethnicity and the number of participants may act as significant confounders in this association.The present meta-analysis evaluated the associations between rs560426 and rs481931 polymorphisms of NSCL/P is a complex congenital anomaly that shows both clinical and genetic heterogeneity and the genetic basis of NSCL/P has remained unclear .ABCA4 in facial cranial morphogenesis, especially since Abca4 null mice do not show cleft palate [ABCA4 polymorphisms (rs560426 and rs481931). In contrast, when ethnicity was considered, among Asians, both the G allele  and GG genotype (compared to AA genotype) were associated with a significantly increased risk of NSCL/P. Next, there is a need to further investigate the possible role of the ABCA4 gene in the etiology of NSCL/P. This need stems from the following observations: First, some studies [ABCA4 gene and the risk of a NSCL/P; second, two GWAS studies [ABCA4 gene with the risk of CL/P; third, the association between the ABCA4 gene and the risk of NSCL/P was observed among at least three different populations, that is to say, among Taiwanese, Honduran and Columbian populations [ABCA4 rs481931 alleles and the status of bilateral and unilateral NSCL/P. Last, the replication of ABCA4 polymorphisms among independent families from various populations revealed that Asian families showed evidence of a higher risk of NSCL/P related to polymorphisms of the ABCA4 gene when compared to the European or American families [Further, there is little biological evidence to support the role of t palate . It turnt palate  and 1.34t palate  increaset palate , 3.0- [2t palate , 1.86- [t palate , and 1.7t palate  elevatedt palate  , and not due to the impact of a single locus. ARHGAP29 is expressed in the palate and lips of mice, its expression depends on Interferon Regulatory Factor 6 (IRF6), and coding variants in ARHGAP29 are related to CL/P [ARHGAP29 is the risk gene at 1p22 [ABCA4 are in linkage disequilibrium with a causal variant included in a neighboring gene, and that associated variants in ABCA4 act as indirect surrogates for a real etiologic variant in individuals with NSCL/P [One study  suggeste to CL/P . Therefo at 1p22 . It follh NSCL/P ,33,37.ABCA4 rs560426 polymorphism and folic acid consumption and the V-maf avian musculoaponeurotic fibrosarcoma oncogene homolog B rs11696257 polymorphism [Besides a purely genetic-based explanation of the etiology of NSCL/P, both gene–gene and gene–environment interactions showed that there was an interaction between the morphism . Transmimorphism .Last, a significantly different rate of the G allele of the rs560426 polymorphism among Asians, but not among Europeans, showed that the genetic variations can be associated with the pathogenesis of NSCL/P, but the role of ethnicity on genetic variations should be taken into consideration.ABCA4 polymorphisms may be in an imbalance linked to polymorphism(s) located in other genes and ABCA4 can be an indirect substitute for NSCL/P etiology. However, the novelty should be balanced against the following limitations. First, the number of studies was rather small. Second, samples were not further investigated for age, sex, and genotyping methods. Third, not all studies reported the Hardy–Weinberg equilibrium. Likewise, fourth, some sources of controls and ethnicities evaluated in the studies were different. However, sensitivity analysis identified the stability of the results, and also the heterogeneity across the studies was minimal.To summarize, the main findings of the present meta-analysis are as follows: the discrepancy between the results of GWAS studies, this meta-analysis, and individual studies suggest that ABCA4 and ARHGAP29 with larger sample sizes in different ethnicities to understand their contribution to the present pattern of results.The findings of the present meta-analysis confirmed that the G allele and GG genotype of the rs560426 polymorphism were significantly associated with the NSCL/P risk in the Asian population, while no such association was observed for the rs481931 polymorphism. There was no association between both polymorphisms (rs560426 and rs481931) and the risk of NSCL/P in the European descent and the mixed ethnicities. There were no differences in the risk of NSCL/P with rs560426 and rs481931 polymorphisms in relation to the source of the controls. Therefore, the ethnicity and the number of participants may act as significant factors in this association. It follows that further studies are needed, focusing on the gene–gene interactions of


SPINK5) polymorphisms  and the risk of atopic dermatitis (AD).This study aimed to investigate the role of serine protease inhibitor Kazal-type 5  with a 95% confidence interval (95% CI) was chosen as the effect size. Egger's linear regression test was enrolled to assess the level of publication bias.Studies associated with P = .0069), co-dominant model , and dominant model . However, no statistically significant difference was found in the recessive model for Asn368Ser and other genetic models for Asp386Asn and Glu420Lys . No significant publication bias was found.Overall, 6 studies met the inclusion criteria for meta-analysis. Significantly statistical differences were calculated between patients with AD and healthy individuals on Asn368Ser polymorphism in the allele model (G vs A: OR = 1.2643, 95% CI = 1.0666–1.4987, SPINK5 Asn368Ser polymorphism may be a risk factor for AD.The  SPINK5 in atopic dermatitis (AD) risk was meta-analyzed.The role of SPINK5 Asn368Ser was significantly associated with AD risk.Polymorphisms of Asp386Asn and Glu420Lys were not associated with AD risk.1 and the prevalence of AD continues to increase in developing countries. Traditionally, AD is often associated with abnormalities in the skin barrier and immune system dysfunction, accompanied by high microbial colonization and a higher susceptibility to skin infection.,4 However, the pathogenesis of AD is not fully understood.Atopic dermatitis (AD), also known as atopic eczema, is considered a chronic inflammatory skin disease. It results in dry, itchy, swollen, and red skin. As of 2018, the point prevalence of adult AD in the overall/treated populations was 4.9%/3.9% in the United States, 3.5%/2.6% in Canada, 4.4%/3.5% in the EU, and 2.1%/1.5% in Japan,SPINK5) is a member of the gene family serine protease inhibitor Kazal-type cluster located on chromosome 5q32, which encode inhibitors of serine proteases. The encoded proteins are mainly distributed in the vaginal epithelium, thymus, vestibular gland, oral mucosa, tonsils, and parathyroid glands, which are mainly involved in the hydrolysis of human growth hormone and skin desquamation. Various mutations in SPINK5 have been identified in patients with AD, and results were widely variable. For example, data from the study by Nishio et al. showed that five missense mutations, such as Asn368Ser, Asp386Asn, and Glu420Lys were associated with AD. However, Jongepier and his colleagues demonstrated that SPINK5 was not associated with atopic phenotypes in individuals ascertained by a proband with asthma. Thus, the association of AD with SPINK5 polymorphisms remains unclear, and results are not conclusive. Therefore, a meta-analysis would be needed to evaluate the role of SPINK5 polymorphisms and the risk of AD.Serine protease inhibitor Kazal-type 5  and AD risk were searched. An odds ratio (OR) with a 95% confidence interval (95% CI) was chosen as the effect size to explore the potential association between SPINK5 polymorphisms and AD risk.In this meta-analysis, previous studies associated with 2The meta-analysis was performed following the guidelines provided by the Preferred Reporting Items for Systematic Reviews and Meta-Analysis Protocols (PRISMA-P). Ethical approval was not necessary since this is a meta-analysis and no patients or animals involved.2.1http://www.ncbi.nlm.nih.gov/pubmed/), Embase (http://www.embase.com), and Cochrane library (http://www.cochranelibrary.com/) with a retrieval deadline of April 22, 2019, based on the predefined search strategy. The keywords and search terms used for all searches were “atopic dermatitis” or “AD” AND “SPINK5” OR “serine protease inhibitor kazal type 5” or “serine protease inhibitor Kazal-type 5” and “SNP” or “Single Nucleotide Polymorphism” or “polymorphism” or “genetic” or “variant”. Finally, in order to enroll more studies, articles of paper literature, and citations were manually screened.Electronic English works of literature were searched from databases, including PubMed  data associated with the genotype and allele frequency of SPINK5 polymorphism  were provided.The present meta-analysis would include the following studies: (1) the research design was a case-control study; (2) the participants in the case group were patients diagnosed with AD, and the participants in the control group were healthy people or hospitalized patients diagnosed without AD; (3) the association between Types of literature would be excluded if they were (1) studies with incomplete data, and statistical analysis could not be performed; (2) reviews, letters, and/or comments. For duplicated publications, only the study with the most complete data, most updated data, or higher Newcastle-Ottawa Scale (NOS) score could be included.2.2.1The authors independently evaluated all relevant articles and extracted relevant data: the first author name, publication year, study region, diagnostic criteria of AD, the detection method of genotype, the same size in the case group and control group, gender, age, and outcomes in each group. Any discrepancies would be resolved by discussion. The genetic polymorphisms mainly included Asn368Ser, Asp386Asn, and Glu420Lys. which was recommended by the Agency for Healthcare Research and Quality (AHRQ) for quality assessment of each case-control study. For example, NOS scores of 0-3, 4-6, and 7-9 represented low, moderate, and high-quality studies, respectively.The NOS was used to assess the quality of enrolled studies,2.3 ORs with their 95% CI of the allele model, co-dominant model, recessive model, and dominant model were calculated in order to assess the relationship between SPINK5 polymorphisms and AD. The heterogeneity was assessed by Dixon's Q-test and I2 test. We defined that significant heterogeneity occurred if P < .05 or I2 > 50%, and then data would be pooled by the random effects model. If P value >.05 or I2 < 50%, data would be pooled by the fixed effect model. The publication bias was evaluated using the Egger's linear regression test with P > .05, indicating no publication bias. Sensitivity analysis was performed by eliminating one study at each defined interval. The results are stable if the outcomes did not change.Data analyses in this study were performed using the R software package 3.12. We initially assessed whether the genotype distribution in the control group was in accordance with Hardy Weinberg Equilibrium (HWE) by using the Chi-square test.33.1,15–19 and were included in the meta-analysis.Figure 3.2 The genotype detection methods were polymerase chain reaction (PCR)-restriction fragment length polymorphism analysis (PCR-RFLP) and/or PCR amplification. Most articles did not report on the gender ration. Table SPINK5 gene polymorphisms  of each included study. It is worth noting that the genotype distribution in the control group of one study  deviated from HWE.The included study characteristics are collected in Table 3.3SPINK5 were evaluated in this study. The allele model , co-dominant model , recessive model , and dominant model  for Asn368Ser, Asp386Asn, and Glu420Lys were evaluated.The associations between AD and the genetic polymorphisms of different genetic models for P < .05, I2 > 50%, Table The heterogeneity test results showed significant heterogeneity among the genetic models of GA vs AA, GG vs AA, GG vs AA+GA, and GG+GA vs AA for Glu420Lys (P = .0069), co-dominant model , and dominant model  indicated significantly statistical differences, while the pooled estimates of the recessive model  were not significantly different . These results demonstrated that the genetic polymorphism of Asn368Ser of SPINK5 was significantly related to AD morbidity and is a risk factor for AD.The pooled estimates for Asn368Ser of the allele model  showed that publication bias among studies was not significant .,22SPINK5 in the epidermis is primarily expressed in the stratum granulosum, where it functions as a protease. Thus, it is important in the cornification of epithelial differentiation and exfoliation.,24 Previous evidence demonstrated that SPINK5 could prevent an influx of pathogens based on the formation of the cornified cell envelope. Evidence from a study by Mocsai et al. showed that skin barrier functions were related to total immunoglobulin E (IgE) levels. Furthermore, Tanei and his colleagues showed that allergic inflammation, mediated by the level of IgE, was crucial in the pathobiology of AD. Recently, Hubiche et al. found an association between SPINK5 E420K polymorphisms and high IgE serum levels. Moreover, several extensive studies support the role of SPINK5 in the development of AD.,18 Thus, the mechanism and the potential role of SPINK5 should be fully elucidated in future studies.It is well known that the skin acts as an essential barrier against pathogens and exogenous agents. Moreover, skin barrier dysfunctions are one of the major factors involved in AD development.SPINK5 Glu420Lys polymorphism genotypes, including GA vs AA, GG vs AA, GG vs AA+GA, and GG+GA vs AA. Recently, it was determined that AD in several ethnic groups displayed variant mutation spots and rates between populations. Moreover, skin barrier dysfunctions introduced by virulence factors could also induce allergic inflammation via innate and adaptive immunity. Thus, limited background information of enrolled patients from a different ethnicity might be possible sources of heterogeneity. Although no significant difference was found in the recessive model for Asn368Ser and other genetic models for Asp386Asn and Glu420Lys in the meta-analysis, further clinical data would also be needed to verify the conclusion.Notably, significant heterogeneity was calculated among data evaluating SPINK5 Asn368Ser polymorphism and risk of AD, whether this SNP could influence gene expression of SPINK5 should be further investigated. Therefore, a study with higher NOS scores and larger sample size would be needed.Furthermore, limitations of this meta-analysis should be noted. Firstly, the enrolled number of patients was small, and subgroup analysis could not be performed. Secondly, the genotyping method was different among enrolled studies, and the sensitivity ability of each method varied, which might lead to false-negatives. Third, though we found the association of SPINK5 Asn368Ser polymorphism as one of the risk factors for patients with AD. Future studies fully elucidating the pathogenic mechanisms involved in the disease are needed.In conclusion, our study supports the role of YLL and HWZ designed the study. YLL received the fund and was a major contributor in drafting the manuscript. HWZ revised the manuscript. YL, WL and XXG searched the references, reviewed the references and extracted the data. SZ performed quality evaluation and statistical analysis. All authors reviewed and approved the final version of the manuscript.