VDR) polymorphisms  and the risk of autoimmune thyroid disease (AITD) had been investigated in previous studies. However, the results of these studies remained controversial. Thus, a meta-analysis was performed to derive a more precise conclusion. All related articles were systematically searched by PubMed, Embase, Google Scholar, and Chinese National Knowledge Infrastructure (CNKI). The pooled odds ratios (ORs) with 95% confidence intervals (CIs) were calculated to assess the strength of association. The overall results indicated that VDR rs731236 and rs2228570 polymorphisms were significantly associated with a reduced risk of AITD. However, a stratification analysis based on clinical types showed that VDR rs731236 and rs2228570 polymorphisms were associated only with a reduced risk of HT. A stratification analysis by ethnicity showed that VDR rs731236 polymorphism was significantly associated with a reduced risk of AITD in Asian and African populations. VDR rs2228570 polymorphism was associated with a reduced risk of AITD in Asian populations. VDR rs1544410 polymorphism was associated with a reduced risk of AITD in European and African populations, but with an increased risk of AITD in Asian populations. VDR rs7975232 polymorphism was significantly associated with an increased risk of AITD in African populations. In conclusion, the present study suggested that VDR rs731236, rs1544410, rs2228570, and rs7975232 polymorphisms were significantly associated with AITD risk. However, more well-designed studies should be performed to verify the current results.The association between vitamin D receptor ( Autoimmune thyroid disease (AITD), mainly including Graves' disease (GD) and Hashimoto's thyroiditis (HT), is an organ-specific autoimmune disease and affects up to 5% of the general population , 2. Alth2D] can promote the differentiation of monocytes and inhibit the maturation of dendritic cells [2D are mediated by its binding to vitamin D receptor (VDR), which belongs to the family of trans-acting transcriptional regulatory factors and is widely expressed in various immune cell subsets, including lymphocytes, macrophages, and several endocrine cells [VDR gene [VDR rs731236, rs1544410, rs2228570, and rs7975232 polymorphisms with AITD risk has been widely reported [Vitamin D is a fat-soluble vitamin and is activated in the liver and kidney . Activatic cells . Furtheric cells . These cne cells . The genVDR gene . Among treported –34. HoweAll related articles were obtained by systematically searching PubMed, Embase, Google Scholar, and Chinese National Knowledge Infrastructure (CNKI). The search keywords were as follows: “vitamin D receptor OR VDR,” “polymorphism OR genetic variation OR genetic variant,” and “autoimmune thyroid disease OR AITD OR thyroid.” There were no limitations on language and year of publication. The last search was updated on August 28, 2017. Furthermore, the references of all related articles were also retrieved to find other eligible studies.VDR polymorphisms  and AITD risk; and (c) available genotype/allele frequencies. In addition, the exclusion criteria were as follows: (a) letters, reviews, and case reports and (b) duplicate publication. If multiple studies had overlapping data, only those with complete data were included.All eligible studies must meet the following inclusion criteria: (a) case-control studies; (b) evaluation of the association between Two authors independently reviewed the related articles and extracted the following data: first author's name, year of publication, region, ethnicity, genotyping methods, the number of cases and controls, and genotype/allele frequency. Any disagreement was resolved by discussion with each other.PHWE < 0.05 was considered as a deviation from HWE. The strength of the association between VDR polymorphisms  and AITD risk was evaluated by the pooled odds ratios (ORs) with 95% confidence intervals (CIs). The significance of the pooled ORs was assessed by the Z test, and PZ < 0.05 was considered statistically significant. The chi-square based Q-test was used to investigate the between-study heterogeneity. If PH < 0.1 indicated the existence of between-study heterogeneity, the random-effect model was used to calculate the pooled ORs; otherwise, the fixed-effect model was applied for the analysis. A sensitivity analysis was conducted by omitting one study each time to estimate the stability of the result. Publication bias was determined by Begg's funnel plot and Egger's test. A symmetric funnel plot and P value of Egger's test more than 0.05 indicated the lack of publication bias. All statistical tests were performed using Review Manager 5.2  and the STATA 12.0 software .Hardy-Weinberg equilibrium (HWE) in the control group of each study was calculated by chi-square goodness-of-fit test, and VDR polymorphisms and AITD risk. Thereinto, there were 15 studies on rs731236, 18 studies on rs1544410, 15 studies on rs2228570, and 16 studies on rs7975232. All studies used polymerase chain reaction-restriction fragment length polymorphism (PCR–RFLP) method for genotyping except for Meng et al.'s study.The flowchart for identifying eligible studies is shown in VDR rs731236 polymorphism and AITD risk is shown in PZ = 0.02; CC versus CT + TT: OR = 0.80, 95% CI: 0.66–0.95, PZ = 0.01). In the stratification analysis based on clinical types, a significant association of VDR rs731236 polymorphism with HT risk was found in the homozygote comparison model . In the subgroup analysis by ethnicity, a significant association was found in Asian  and African populations . The pooled analysis based on PHWE > 0.05 showed a significant association in the allele comparison model .The association between VDR rs1544410 polymorphism with AITD risk is presented in VDR rs1544410 polymorphism was associated with a reduced risk of AITD in European  and African  populations and with an increased risk of AITD in Asian populations . The pooled analysis based on PHWE > 0.05 showed that VDR rs1544410 polymorphism was associated with a reduced risk of AITD .The association of VDR rs2228570 polymorphism and AITD risk is shown in PZ = 0.02; TT + CT versus CC: OR = 0.71, 95% CI: 0.54–0.93, PZ < 0.001; T versus C: OR = 0.80, 95% CI: 0.68–0.95, PZ = 0.01). A further stratification analysis by clinical types showed a significant association in HT  but not in GD. A subgroup analysis based on ethnicity showed a significant association in Asian populations , but not in European populations. A stratification analysis by PHWE value showed a significant association in studies of PH ≤ 0.05  but not in studies of PH > 0.05.The association between VDR rs7975232 polymorphism, no significant association was observed in the overall analysis and stratification analysis by clinical types and PHWE .For nd PHWE . HoweverP values of Egger's test were more than 0.05, indicating the lack of publication bias  in the VDR gene could affect the expression of VDR [VDR gene was associated with higher VDR mRNA copy numbers [VDR gene could affect VDR gene expression by modulating mRNA stability [VDR rs7975232 polymorphism was significantly associated with GD risk in Chinese populations [VDR rs2228570 polymorphism and HT risk in Serbian populations [VDR rs731236, rs7975232, and rs1544410 polymorphisms were associated with GD susceptibility in Eastern Croatian populations [VDR rs731236 and rs2228570 polymorphisms were significantly associated with HT risk in a Turkish population [VDR rs1544410 and rs731236 polymorphisms were significantly associated with a reduced risk of AITD, while VDR rs7975232 and rs2228570 polymorphisms were not associated with AITD risk [VDR rs731236 and rs2228570 polymorphisms were significantly associated with reduced risk of AITD. A further stratification analysis based on clinical types showed that VDR rs731236 and rs2228570 polymorphisms were associated only with reduced risk of HT. In a stratification analysis based on ethnicity, VDR rs731236 polymorphism was associated with a reduced risk of AITD in Asian and African populations but not in European populations. VDR rs2228570 polymorphism was associated with a reduced risk of AITD only in Asian populations. It was worthy to note that VDR rs1544410 polymorphism was associated with a reduced risk of AITD in European and African populations but has an increased risk of AITD in Asian populations. A stratification analysis by PHWE values showed that the significant association of VDR rs2228570 polymorphism with a reduced risk of AITD was observed only in studies with PHWE ≤ 0.05, which indicated that the effect of VDR rs2228570 polymorphism on AITD risk needed to be interpreted cautiously.As an immune modulator, vitamin D is involved in the onset and development of AITD , 36. Lown of VDR , 39. For numbers . Uitterlulations , 15. Djuulations . Stefanipulation . Howeverpulation , 40, 41.ITD risk . Due to ITD risk , 15, 16.Heterogeneity was observed in the current meta-analysis. We tried to investigate the sources of heterogeneity by a stratification analysis based on clinical types, ethnicity, and HWE, but the investigation results were not satisfactory and could not provide a reasonable explanation for the sources of heterogeneity. In view of factors affecting vitamin D levels and methodological issues, heterogeneity may result from differences in economic and public health indexes among different countries, variations in environment and climate, age and gender mismatch in published studies, and variations in diagnostic criteria for GD/HT.VDR polymorphisms with AITD risk, such associations had a small influence on the occurrence of AITD. In addition, several limitations impeding accurate assessment should be noted. Firstly, raw data such as gender, age, living style, and drug consumption could not be obtained from all included studies. Secondly, the number of studies was small in the subgroup analysis, especially in African populations. Last but not least, the present analysis did not consider the gene-gene and gene-environment interactions.Although the current results showed the statistically significant associations of VDR rs731236, rs1544410, rs2228570, and rs7975232 polymorphisms were significantly associated with AITD risk. However, more well-designed studies, especially studies on African populations, should be performed to verify the results.In conclusion, the present study suggested that


P = 0.001 in allele model; OR = 0.769, 95%CI: 0.650-0.909, P = 0.002 in recessive model; OR = 0.505, 95% CI: 0.352-0.726, P = 0.000 in dominant model; for 11536889: OR = 0.927, 95% CI: 0.872–0.984, P = 0.013 in allele model; OR = 0.926, 95% CI: 0.862–0.944,P = 0.034 in recessive model. In terms of subgroup analyses sorted by ethnicity, only polymorphism of rs4986791 had a significant influence on decrease of cancer risk among both Caucasian and Asian populations. The findings suggested that TLR4 polymorphisms may serve as a genetic risk factor for cancers.Previous studies have explored the association between toll-like receptor 4 (TLR4) polymorphisms and risk of various cancers, but the results remained controversial. To obtain an assessment of the effect of TLR4 polymorphisms  on cancer risk, fifty-five articles (containing 20107 cases and 28244 controls) were recruited for meta-analysis. Our result indicated that two Single Nucleotide Polymorphisms (SNP) in TLR4 were associated with decreased cancer risk for rs4986791: OR = 0.764, 95% CI: 0.652-0.894,  Toll-like receptors (TLRs) were important innate immune molecules which were used for specific recognition to adjust the adaptive immune response by identifying the pathogen-associated molecular patterns (PAMPs). In addition, TLRs stimulated the immune system and promoted the development of cancer due to its potential chronic inflammation –3. In reRecent studies indicated that TLR4 expressed on many cells as the bridge between the natural immunity and acquired immunity was able to produce proinflammatory factors and chemotactic factors after combining with ligands such as lipopolysaccharide(LPS) and teichoic acid . MoreoveNumerous publications demonstrated that the existence of single nucleotide polymorphism (SNPs) made it possible for affecting TLR4 signaling, which was responsible for hyporesponsiveness and infection of fungus, virus and bacteria and the expression of TLR4 had a close association with the development of breast cancer –14, lungIn the beginning, there were 166 relevant articles searched mainly from PubMed, of which 26 duplicates were excluded. And 43 articles were eliminated because of their irrelevance with cancer. After getting rid of studies focusing on other gene instead of TLR4, meta-analysis, studies lacking comparison between case and control and studies having few valuable data, 55 publications were eventually included. As elaborated in Figure TLR4 rs4986790 polymorphism and cancer risk. Before the analysis, we tested the sensitivity of this SNP and found an article affecting the sensitivity greatly through leave-one-out method. Therefore, this article written by Hold GL1  about gaP = 0.001; recessive model: OR = 0.769, 95% CI: 0.650–0.909, P = 0.002; dominant model: OR = 0.505, 95% CI: 0.352–0.726, P = 0.000), Caucasian population  and Asian population  in three genetic models.TLR4 rs4986791 polymorphism and cancer risk. To investigate the relation between the polymorphism and cancer risk, we searched 27 studies for useful data and 4416 cases and 7379 controls were analyzed. Table P = 0.013; recessive model: OR = 0.926, 95% CI: 0.862–0.944, P = 0.034). The similar association was also found for rs11536889 in Asian population under allele model  and dominant model  illustrated in Table TLR4 rs11536889 polymorphism and cancer risk. Taking 14 articles, containing 6586 cases and 9527 controls, into consideration, we noticed a connection between rs11536889 polymorphism and decreased risk of cancer in overall population under two genetic models . We performed logistical meta-regression to evaluate the source of heterogeneity among all studies and found that ethnicity had a significant influence on heterogeneity, while cancer type, number of alleles, sample size and genotyping methods could not greatly influence the initial heterogeneity. According to Cochrane handbook [I2 is lower than 50% is acceptable confirming the credibility of our study.As shown in Tables P = 0.496 for rs4986790, after the article of Hold GL was excluded from this study; P = 0.297 for rs4986791; P = 0.827 for rs11536889, respectively). The leave-one-out sensitivity analysis was also performed to examine the effects of individual article on the pooled ORs. The result, illustrated in Figure The publication bias of included literatures was evaluated through the Begg’s test under allele model. As displayed in Figure In this study, we conducted a meta-analysis of 55 independent articles concerning the association between TLR4 polymorphisms and cancer risk. As a result, rs4986791 polymorphism had a significant association with decreased cancer risk under 3 genetic model. And similar association was also found between rs11536889 and risk of cancer in overall and Asian populations.TLR4 gene, located in chromosome 9q32-q33, was an important receptor of lipopolysaccharide and played a key role in immune responses thorough activation of nuclear factor kappa-light-chain-enhancer , was assAs for stratified analyses by ethnicity, polymorphism of rs4986791 had a prominent effect on decreased cancer risk of Caucasian and Asian. Also, decreased cancer risk among Asian population was associated with rs11536889 polymorphism.In terms of subgroup analyses classified by race and cancer type, we found that Caucasian female-specific cancer had a conspicuous association with polymorphisms of rs4986790 and rs4986791 and Asian digestive cancer was significantly influenced by rs4986791 polymorphism. There was no evidence for another association between these SNPs and cancer risk.In summary, this meta-analysis illustrated polymorphisms of TLR4 rs4986791 and rs11536889 might make contributions to slow the formation and development of cancer, whereas rs4986790 was not strongly associated with cancer risk. Due to the limit of sample size of African population in this study, there is still a requirement of researches including more data for getting insight into association between TLR4 polymorphisms and cancer risk.st, 2017 ). The key words used in searching were ‘TLR4 or toll-like receptor’ and ‘cancer’ and ‘SNP or polymorphism’. As illustrated in Figure To investigate the association between TLR4 polymorphisms and cancer risk, we searched PubMed and some other databases for relevant articles, ranging from 2000 to 2017  the studies focused on TLR4 and cancer risk;(b) each publication was case-control study and had valuable data for calculating genotype counts;(c) the cancers were classified by the experienced pathologist, according to the World Health Organization (WHO) criteria; (d) the articles were written in English or Chinese. The following information is shown in our study: first author, journal, year of publication, ethnicity, cancer type, sample size, genotype and allele frequency.P < 0.05 was regard a significant departure from HWE. Allele model, recessive model and dominant model were compared to evaluate the association between TLR4 polymorphisms  and the risk of cancer.In this study, we used the software Stata 12.0  for meta-analysis of the association between TLR4 polymorphisms and cancer risk, which was estimated by odds ratio (OR) and 95% confidence intervals (CI). In addition, we conducted stratification analysis by ethnicity  and cancer type  to understand the effects of them on the association. Hardy-Weinberg equilibrium (HWE) was estimated for each study by Pearson Chi-square test in control group and Q-test and I2 statistics were selected. When the heterogeneity was obvious (Ph> 0.05), the mixed effects model (the Mantel-Haenszel method) was used for the summary OR value. Otherwise, the random effects model (the DerSimonian and Laird method) took the place of it.We adopted Begg’s test and Funnel plots for evaluation of publication bias of this study and exclusion of one article at a time leading to new pooled ORs, which showed the influence of individual article on the overall consequence. Moreover, to examine heterogeneity among all the articles, the


Nine publications were included in the final analyses, consisting of 18 independent cohorts with 22 risk estimates . Results of the dose-response analyses showed a nonlinear inverse relationship between BMI and NMSC , which persisted when limiting to the studies with adjustment for important potential confounders including sun exposure and sensitivity factors. The risk estimates were very similar for squamous cell carcinoma and basal cell carcinoma. Sex appeared a source of heterogeneity (P-difference = 0.06), with a weaker, but still significant inverse association in men than in women. This dose-response meta-analysis suggests a nonlinear inverse association between BMI and NMSC.Prospective epidemiologic studies that investigated the association between body mass index (BMI) and non-melanoma skin cancer (NMSC) yielded inconsistent findings. A dose-response meta-analysis was conducted to quantitatively summarize the evidence. PubMed and Embase databases were searched for relevant studies. Study-specific relative risk (RR) and 95% confidence interval (CI) for an increase in BMI of 5 kg/m Non-melanoma skin cancer (NMSC), generally referring to squamous cell carcinoma (SCC) and basal cell carcinoma (BCC), ranks the most frequently diagnosed cancer worldwide. Unlike other cancers for which the incidence rates has become stabilized or declined, the incidence of NMSC is still increasing, in particular among younger populations15Obesity, as measured by body mass index (BMI), is an established risk factor for various chronic diseases such as type 2 diabetes, cardiovascular disease, and cancers at various sites, as well as all-cause mortality51011121314151617101116121314Potentially relevant publications were searched on PubMed and Embase databases from inception to 30 June, 2016. The detailed literature search strategy is reported in 2), or obesity status measured by BMI; (3) the outcome of interest was NMSC or its subtype; and (4) relative risks (RRs) and 95% confidence intervals (CIs) were reported for continuous BMI , or for at least 3 quantitative categories of BMI. When the same study was reported in overlapping publications, the one containing the largest number of NMSCs was included.Studies were eligible for inclusion if they met the following criteria: (1) the study design was prospective ; (2) the exposure of interest was BMI  duration of follow-up, sex and age of participants, number of cases and participants, methods for BMI measurement, categories of BMI with corresponding RR and 95% CI, and potential confounders accounted for in the statistical analyses. Two authors (ZD and LG) independently conducted literature search, study selection, and data abstraction, with any disagreement resolved by consensus. Instead of assessing methodologic quality of the primary studies by a quality score, we investigated whether the summary risk estimates were significantly influenced by some study and population characteristics  which may be indicators of study quality2 increase in BMI. When only results for categorized BMI were reported, the method of generalized least squares trend estimation proposed by Greenland and Longneckeret al.In this meta-analysis, we used the most fully adjusted risk estimates, and reported summary RR of NMSC for each 5 kg/mP value for non-linearity was calculated by testing the null hypothesis that the coefficient of the second spline is equal to 0. Heterogeneity among studies was assessed with the Q and I2 statisticsQ statistic, a P < 0.1 was considered significant heterogeneity. Potential publication bias was assessed by a combination of Begg rank correlation test, Egger linear regression test, and Begg’s funnel plot25To explore potential sources of heterogeneity, subgroup analyses were performed by geographic area, sex of participants, NMSC subtype (SCC and BCC), methods for BMI measurement, and adjustment for important confounders. Further sex-specific analyses were conducted for SCC and BCC. A potential nonlinear association between BMI and risk of NMSC were investigated using restricted cubic splines with three knots at percentiles 10%, 50%, and 90% of the distributionThe flow diagram of study selection is reported in Finally, 9 publications10111213141516171112151617101314111612122 increase in BMI, with considerable heterogeneity  , which was supported by a visual inspection of the Begg funnel plot that showed no obvious asymmetry  for each 5 kg/m= 71.2%) . There w1.2%) . There also a suggestion of an effect modification by sex (Pdifference = 0.06), with a weaker but still significant association in men than in women (Pdifference = 0.004), but only 3 reports were included in the group with adjustment. Of note, the association did not differ according to adjustment or not for several well known risk factors for NMSC such as family history and indicators of sun exposure and sun sensitivity. Excluding the 3 studies (7 reports) where participants were generated from clinical prevention trials, the summary RR was 0.86 (95% CI: 0.83–0.90), with substantial heterogeneity (I2 = 77.0%).Results of subgroup analysis performed to explore potential sources of heterogeneity according to pre-specified factors are reported in in women . This sein women . There wIn this dose-response meta-analysis of prospective studies involving 18 independent cohorts with nearly 1 million participants and over 50,000 NMSC cases, we found that BMI was nonlinearly, inversely associated with risk of NMSC. The association was similar for SCC and BCC, and persisted when limiting to the studies with adjustment for important potential confounders including sun exposure and sun sensitivity measures. We also found that the inverse association in women was more evident than in men.2 was non-significantly inversely associated with NMSC risk Potential biological mechanisms under the observed inverse association between BMI and NMSC risk are poorly understood. Some investigators attributed this association to the residual confounding by sun exposure16172930et al.35Gerstenblith Evidence from case-control studies on BMI and NMSC has been limited, and some also suggested a sex-specific difference. In a small, hospital-based case-control study of 77 NMSCThis meta-analysis presents several strengths. Limiting our analyses to prospective studies eliminated reverse causality because BMI was determined before the diagnosis of cancer. The large number of cases strengthened the statistical power to detect a weak association. Our linear and nonlinear dose-response analyses enabled us to better understand the shape of the association. Moreover, the stability of our results was supported by the stratified analyses according to various important study characteristics. Heterogeneity and publication bias are two important concerns for a meta-analysis of published literature, because different studies would vary by population and methodological characteristics, and small studies with null results have less chance to be published. However, with commonly used statistical methods, we found little evidence of publication bias. We were also able to identify potential sources of heterogeneity by performing meta-regression analyses.However, this study is not without limitations. As a meta-analysis of observational studies, the problem of potential confounders is the first consideration. Although the summary risk estimates were not altered when restricting to the studies with sun exposure and sensitivity adjustment, residual confounding arising from measurement errors of sun exposure  remained. Original studies generally used self-reported questionnaires to collect lifestyle information, and the way in which they asked about the duration and density of the exposure varied across different questionnaires; moreover, other problems such as the use of sunscreen and the degree to which clothing covers the body surface were less or not considered in these cohorts. Therefore, future prospective studies with more accurate measures of factors related to both BMI and sun exposure, and those studies using a Mendelian randomization design are needed to further address the impacts of these confounders on the association.In summary, findings from this meta-analysis of prospective studies suggest that BMI is nonlinearly and inversely association with risk of NMSC. The inverse association is similar with SCC and BCC, but it appears stronger in women than in men. Potential mechanism underlying the association remains to be determined. Despite that we and others have observed lower risks of specific cancers .Publisher's note: Springer Nature remains neutral with regard to jurisdictional claims in published maps and institutional affiliations.


The angiotensin-converting enzyme (ACE) I/D polymorphism has been reported tobe associated with Kawasaki disease (KD), but studies to date presentconflicting results.The aim of this study is to derive a more precise estimation of theassociation between the ACE I/D polymorphism and KD risk.PubMed, EMBASE, CNKI and Wangfang databases were retrievaled to identify forrelevant studies from inception to May 2017. Pooled odds ratios (OR) with95% confidence intervals (CI) were calculated using Stata 12.0 software.A total of 6 case-control studies comprising 634 patients and 458 controlswere included in the meta-analysis, and we found a significant associationbetween the ACE I/D polymorphism and KD risk .When stratified by sample size>200, this polymorphism is associatedwith an increased the risk of KD.The I/D polymorphism in the ACE gene may be associated with susceptibility toKD. Kawasaki disease (KD) is an self-limited vasculitis that mainly affects youngchildrenThe renin-angiotensin system (RAS) has been implicated in modulating blood pressureand homeostasis of the cardiovascular systemMany studies have investigated the relationship between ACE I/D polymorphism and KD.The inconsistency of these results may have resulted from inadequate statisticalpower owing to small sample size and eco-geographical differences. Meta-analysis mayovercome these limitations of individual researchComputer searches of PubMed, EMBASE, CNKI and Wangfang databases were performedvia the following key words: “ACE gene”, “Kawasakidisease/KD”, “I/D”, “single nucleotidepolymorphism” and “genetic polymorphism”. Only humanstudies were selected. Additional articles were identified by a manual searchingof the references of in the related original studies.Articles included in the meta-analysis met the following inclusion criteria:1)relevant case-control studies of KD cases and healthy controls; 2) articles onthe relation of the ACE I/D polymorphism and susceptibility to KD and 3) studiesthat included sufficient genotype information for extraction. Exclusion criteriawere as follows: 1) not case-control studies; 2) case reports, reviews, ormeta-analysis; 3) studies that were based on incomplete raw data.The collected data included the first author's surname, publication date,country of origin, ethnicity, the number of cases and controls, the genotypefrequency of ACE I/D polymorphism and deviation from Hardy-Weinberg Equilibrium(HWE) of the control group.Fisher's exact test was used to test HWE for distributions of genotypes amongcontrols. The strength of the correlation between ACE I/D polymorphism andsusceptibility to KD was estimated by odds ratio (OR) and 95% confidence interval(95%CI) as follows: D vs I, a homozygote comparison (DD vs II), a heterozygotecomparison (DI vs II), a dominant model (DD+DI vs II) and a recessive model (II+DIvs DD) between groups. The heterogeneity among these articles was checked via the I2test. When I2 > 50% indicated heterogeneity across studies, the randomeffects model was used, otherwise the fixed effects model was performed. Thesensitivity analysis was performed by used via omitting each individual article, andan individual article was suspected of excessive sensitivity if the point estimateof its omitted analysis was outside the 95% CI of the pooled analysis. To assess thepotential publication bias, Begg's and Egger's tests were performed.All statistical tests were performed with STATA .The database search yielded 137 publications, of which both of the reviewersconsidered 10 to be potentially eligible. We excluded 4 of the articles duringthe second phase of the inclusion process. The remaining 6 articles wereincluded in the combined analysisA summary of the meta-analysis findings of the association between ACE I/Dpolymorphism and KD risk is shown in To evaluate the effect of a single article on the final result, we used asensitivity analysis via removing one study at a time. Ultimately the pooledresults hardly changed after removal of each study, suggesting that our resultswere robust .Begg's and Egger's tests were used to assess the publication biasfor ACE I/D polymorphism. The shape of the funnel plot did not reveal anyevidence of obvious asymmetry, suggesting no evidence of publication bias forACE I/D polymorphism .Although the morbidity is highest in Asians, KD is a major cause of acquired heartdisease throughout the worldIn this meta-analysis, we addressed the association between ACE I/D polymorphism andsusceptibility to KD. Our results indicated that the ACE I/D polymorphism wassignificantly associated with the risk of KD. Nevertheless, considering that otherpotential factors might influence the final result, we conducted sub-group analysis.In a stratified analysis by sample size, pooled results showed significantassociation with sample size>200 but not with sample size≤200,suggesting that there was no small-study bias in this meta-analysis. The results ofour study differ from a previous meta-analysisSeveral limitations should be acknowledged of the current study. First, in the pooledanalysis, we found that the ACE I/D polymorphism was significantly associated withthe risk of KD in studies with PHWE<0.05. The data indicated that selectionbias or genotyping error may have affected the merged results. Second, only the ACEI/D polymorphism was analyzed in this meta-analysis. Further analysis should clarifythe association of other polymorphisms in the RAS genes, such as the AGT M235T andT174M polymorphisms. Third, we were unable to include unpublished studies, whichmight affect the publication bias. Additionally, there is a lack of information forthe other population outside Asia. Therefore, the results of the current study arenot comprehensive.Our pooled data showed a significant association between the ACE I/D polymorphism andthe risk of KD. Due to the defect limitations of the included research, futurelarge-scale investigations with appropriate design are required to confirmassociation.


In summary, statin use is associated with a decreased incidence rate of cirrhosis and is most pronounced in Eastern countries but also in Western countries.Hepatitis B and hepatitis C are leading causes of chronic liver disease, particularly cirrhosis. Recently, several studies have observed that statins have an inverse relationship with cirrhosis in hepatitis B or C patients. However, no published meta-analysis studied the protective effect of statins on cirrhosis. Thus, we conducted a systematic review and meta-analysis of published observational studies to better understand the relationship between statins and the risk of cirrhosis. Relevant studies were identified by searching PubMed, EMBASE, and ISI Web of Science for articles published before April 2017. The Newcastle-Ottawa Scale was used to evaluate the quality of the included studies. Six cohort studies, including 38951 cases of cirrhosis in 263573 patients with hepatitis B or C, were identified to investigate the relationship between statins and the risk of cirrhosis. The Newcastle-Ottawa Scale scores for the included studies ranged from 6 to 9, with four high-quality studies and only two of medium quality. The use of statins was associated with a significant 42% reduction in the risk of cirrhosis, without obvious heterogeneity. In addition, this protective effect was more obvious in Asian countries. Moreover, dose-response analysis suggested each additional 50 cumulative defined daily doses (cDDD) of statins decreases the risk of cirrhosis by 11% (odds ratio [OR] = 0.89, 95% confidence interval [CI] = 0.86–0.93,  Globally, approximately 130–170 million people have hepatitis C virus infection, which is equivalent to 2% of the world's population , 5; MoreStatins are a major cholesterol-lowering drug that has been used to prevent and treat various cardiovascular diseases. Recently, other potential benefits of statins have attracted increasing attention worldwide. For example, studies have indicated that statins can decrease the risk of some cancers, including prostate, colorectal, lung, breast and liver cancers –15. In aFigure The main characteristics of the included studies are listed in Table 2 = 31.9%; p = 0.142). These results are shown in Figure Six cohort studies –26 were 2 = 58.0%)  Table .p = 0.001) .No testing for funnel plot asymmetry was performed because of the restricted number of included studies in the analysis  was performed for a more comprehensive assessment in our current meta-analysis. The cumulative Z-curve has crossed the monitoring boundaries already, demonstrating that our results were based on sufficient evidence. This finding revealed statin use were strongly associated with cirrhosis risk Figure .Although published meta-analyses suggested a protective association between statin use and risk of hepatocellular carcinoma –29, no pOur study only demonstrated an association between the use of statins and a reduced risk of cirrhosis; the data cannot establish a causative role for statins in this regard. However, if such a causative role is present, possible mechanisms could be the following. First, statins can inhibit the expression of fibrogenic cytokines such as connective tissue growth factor (CTGF), platelet-derived growth factor (PDGF) and transforming growth factor-b (TGFb), which play a critical role in the development of cirrhosis –32. SecoOur study has several strengths. First, it is the first to explore the dose–response relationship between statin use and the risk of cirrhosis in hepatitis B or C patients. Therefore, this may provide insight into the relationship between the use of statins and the risk of cirrhosis. Second, we performed subgroup and sensitivity analyses to determine which factors affect the risk. Third, most of the studies included in our meta-analysis were of high quality. Additionally, the studies included in the meta-analysis were cohort studies, which are not prone to generate recall and selection biases. All of these characteristics make the conclusions of our study more convincing.There are several limitations that we must consider. First, there are many types of statins, including atorvastatin, cerivastatin, fluvastatin, lovastatin, pravastatin, rosuvastatin, simvastatin, and pitavastatin. However, we failed to obtain information about the use of other types of statin, which might have an influence on the final results. Second, the outcome that we observed was an association, which is subject to confounding bias. Although we considered several adjustment factors, there remain many potential adjustment factors that are unknown, such as the cholesterol level, the triglyceride level, or other over-the-counter drug use, which are closely related to the development of cirrhosis. In addition, we failed to obtain information about antiviral treatment in patients with HBV or HCV, which could have influenced the development of cirrhosis. Third, only 6 studies were included in our article, this puts the meta-analysis at high risk of publication bias. Additionally, Studies with different outcomes are combined which leads to the concern of heterogeneity in the meta-analysis. For example, in Simon 2015 , the outIn summary, our meta-analysis indicated that the use of statins is associated with a 42% lower risk of cirrhosis in hepatitis B or C patients. Moreover, this protective effect is more obvious in Asian countries. However, more prospective studies and basic research are still urgently needed to further validate the association between the use of statins and the risk of cirrhosis as well as the potential mechanisms.We searched published reports in the PubMed, EMBASE and Web of Science using the following keywords: (“statin* OR atorvastatin OR cerivastatin OR fluvastatin OR lovastatin OR pravastatin OR rosuvastatin OR simvastatin OR pitavastatin”) and (“cirrh* OR fibro*”). The initially relevant studies were identified up to April 2017, with no restrictions on the language of publication. We have extra data through contacting the original researchers.The eligibility criteria were as follows: study design (case control or cohort); exposure factor statin and outcome cirrhosis or fibrosis; and odds ratio (OR)/risk ratio (RR) values and corresponding 95% confidence intervals (CIs) for different categories of statin use available or sufficient information provided to enable the calculation of these variables. If two studies reported the same data, we selected the study with the larger sample.Two researchers (JPX and YQW) independently extracted the required information from the selected reports in a standardized manner. We collected the following information from each article: year of publication, first author's name and country of origin, study design (case control or cohort), number of participants , duration of follow-up, sources of controls, comparison of exposure levels, potential adjusted confounding variables, OR/RR values and 95% CIs for different categories of statin use. To assess the dose–response, we also collected the number of case and person-years for each category of statin use.We used the Newcastle-Ottawa Scale  to evaluWe assessed the relationship between statins and cirrhosis using OR/RR values and the corresponding 95% CIs. When the results provided were for multiple groups with the use of statins with OR or RR values and corresponding 95% confidence intervals, we combined them to obtain a single OR/RR value and corresponding 95% CI . We treap value for the non-linear dose-response relationship was calculated by testing whether the coefficient of both the second and the third spline was zero [To enable the meta-analysis of the dose–response, we extracted the number of cases and person-years and RRs with variance estimates for at least three quantitative exposure categories from each study. If the studies did not provide these data, we required sufficient information to calculate them. For dose–response analysis, we used the midpoint of statin use in each category as the dose of statin use. If the highest category was open ended, we set the midpoint of the category at 1.5 times the lower boundary; if the lowest category was open ended, we set the lowest boundary at zero . We obtawas zero . Greenlawas zero , 42, andwas zero , 44.2 to assess heterogeneity between studies and defined low, medium, and high heterogeneity as 25%, 50%, and 75%, respectively [We used Iectively . If p waectively , Egger'sectively  and funnectively .We also performed subgroup analyses by geographic area, number of cases, study quality, and whether alcohol intake, smoking, body mass index or diabetes was adjusted for in the models. Sensitivity analyses were performed by changing the pooling model (random-effects model or fixed-effects model) and excluding studies that were not eligible for dose–response analysis. Sensitivity analysis was also performed to assess the effect of every study on the summarized estimate by sequentially excluding one study in one turn.All statistical analyses were performed using STATA version 12.0 (Stata).Random error can mislead results in meta-analyses. The random error risk may increase remarkably because of multiple looks on accumulating evidence when new trials emerge . To obta


CFHRs) gene deficiency, complement factor H (CFH) autoantibodies, and atypical hemolytic uremic syndrome (aHUS) susceptibility. EMBASE, PubMed, and the ISI Web of Science databases were searched for all eligible studies on the relationship among CFHRs deficiency, anti-FH autoantibodies, and aHUS risk. Eight case-control studies with 927 cases and 1182 controls were included in this study. CFHR1 deficiency was significantly associated with an increased risk of aHUS (odds ratio (OR) = 3.61, 95% confidence interval (95% CI), 1.96, 6.63, p < 0.001), while no association was demonstrated in individuals with only CFHR1/R3 deficiency . Moreover, a more significant correlation was observed in people with both FH-anti autoantibodies and CFHR1 deficiency  in contrast to those with only CFHR1 deficiency. In addition, the results were essentially consistent among subgroups stratified by study quality, ethnicity, and gene detection methods. The present meta-analysis indicated that CFHR1 deletion was significantly associated with the risk of aHUS, particularly when combined with anti-FH autoantibodies, indicating that potential interactions among CFHR1 deficiency and anti-FH autoantibodies might impact the risk of aHUS.To evaluate the association among complement factor H-related ( Atypical, non-diarrhea-associated hemolytic uremic syndrome (aHUS) is an aggressive, fatal systemic disease, which is characterized by thrombocytopenia, microangiopathic hemolysis, and acute renal injury . The maiCFHRs (CFHR1-5) encoding FHR proteins and manifesting high degrees of sequence identity with CFH indicated that mutations in CFHRs may be partly responsible for the function abnormalities of CFH [CFHR deficiency and the pathogenesis of aHUS [CFHR1 deficiency, others have failed to find this correlation [CFHR1 deficiency took part in the formation of CFH dysfunction, indicating that both autoantibodies and CFHR1 absence might participate in the occurrence of aHUS. However, there still remains a need to attach importance to the effect of interactions among autoantibodies of the complement factors and CAP regulators on the development of aHUS [The dysregulation of the alternative pathway of the complement cascade has been confirmed to play a pivotal role in the development of aHUS . Mutatios of CFH . Several of aHUS ,13. Althrelation ,15. Furt of aHUS ,17. CFHRs deficiency, anti-FH autoantibodies, and aHUS risk.Therefore, we conducted a meta-analysis of all eligible studies to estimate the association among CFHRs” or “complement factor H-related protein” or “FH-related proteins” or “anti-FH” or “factor H antibodies” or “anti-complement factor H autoantibody”) and . There were no language restrictions. In addition, we manually searched the reference lists and pertinent reviews of retrieved articles for published or unpublished data. When datasets were incomplete for the required data, ongoing investigations or related experts were contacted for more information.A systematic literature search of PubMed, EMBASE, and the ISI Web of Science databases was performed up to March 2016. The following search terms and key words were used:  the original article was published as a cohort, case-control, or cross-sectional study; (3) the article provided the odds ratio (OR), relative risk (RR), or hazard ratio (HR) with a 95% confidence interval (95% CI) or sufficient data was provided to calculate it. If the articles reported on the same or overlapping data, only the article with the most updated data was included in this study. Three investigators  identified the eligible studies independently. Any differences in opinion were resolved through consensus with a third author (Le Ma) for adjudication.An initial screen of abstracts and titles was performed to identify all potential retrieved articles. Thereafter the full text was reviewed to determine whether the articles were suitable for this systematic review. Articles were eligible for inclusion in this study if they fulfilled the following criteria: (1) the study objective was to evaluate the relationship among The following data were extracted using a standardized data extraction form that included: first author’s name, publication year, country of origin, study design, characteristics of the study population , gene and protein detection methods. If a study provided several risk estimates, only the most completely adjusted estimate was extracted. The quality of studies was evaluated based on the Newcastle-Ottawa Quality Assessment Scale (NOS) . Each arCFHRs, anti-FH autoantibodies, and aHUS, using a fixed effects model or a random effects model according to the heterogeneity. Heterogeneity between studies was examined by calculating Cochran’s Q statistic and the I2 test . When significant statistical heterogeneity was detected, a stratified analysis was conducted to explore the potential source of heterogeneity. The variables included in the stratified analyses were study quality, ethnicity, and gene detection methods. In addition, sensitivity analyses were performed to evaluate the effect of each individual study by excluding each study from the meta-analysis. Egger’s weighted regression method, Begg’s rank correlation method, and visual inspection of funnel plots were used to assess the publication bias [The relative risk (RR) was considered as a measure to estimate the strength of the association across studies in principle. For case-control studies, relative odds ratios (OR) were considered as a surrogate estimate of the corresponding RR. Because the absolute risk of aHUS is low, the OR could approximate the RR. If articles did not report RR, we used the number of events in each group to calculate the RR for each study and used them for pooling analyses. In this meta-analysis, pooled RR and 95% CI were used to estimate the strength of the association among ion bias ,20. All A total of 1145 articles were retrieved through comprehensive searching. After excluding duplicates and irrelevant references, 39 records were left for full text and data effectiveness review, and eight articles (nine studies) were included in this meta-analysis ,21,22,23CFHR deficiency through multiplex ligation-dependent probe amplification. The other three studies used polymerase chain reaction, comparative genomic hybridization, or Western blot, respectively. Six of the eight studies detected anti-FH autoantibodies in serum samples using enzyme-linked immunosorbent assay. Furthermore, quality assessment revealed that six studies scored ≥5 and were classified as high quality.The characteristics of the included studies are presented in CFHR1 deficiency and the risk of aHUS was evaluated in seven studies, including 832 cases and 1028 controls. Among the studies, five reported a statistically significant relationship. There was significant heterogeneity  between the data sets, and the random-effects model was used to calculate the pooled OR. The pooled results showed that CFHR1 deficient individuals had a higher susceptibility to aHUS by approximately 2.6-fold in comparison with no-deficiency patients .The relationship between CFHR1/R3 deficiency (without anti-FH autoantibodies) and aHUS risk, with a total of 477 cases and 436 controls. Three reported a slight association of such deficiency with aHUS incidence. Findings from the current analysis showed that there was no significant association among CFHR1/R3 absence and the risk of aHUS . Subsequently, we also performed stratified analysis by study quality, ethnicity, and gene detection methods. The result showed that these factors did not alter the association direction and magnitude of the overall effect. Moreover, no publication bias was detected and the funnel plots were symmetric upon visual inspection (p = 0.46), which is confirmed by the Egger’s test (p = 0.35).Five studies reported data on the association between only CFHR1 deficiency on the risk of aHUS in five studies, which included 545 cases and 377 controls. Of the included studies, three showed that both CFHR1 deficiency and autoantibody had a positive relationship with aHUS risk. There was no evidence of statistical heterogeneity among the eligible studies , and the fixed-effects model showed that individuals with CFHR1 deletion and anti-FH autoantibodies had an obviously increased risk of aHUS  nor the Begg’s test (p = 0.46) revealed significant publication bias.We subsequently evaluated the effect of anti-FH autoantibodies with CFHR1 was significantly associated with aHUS. Additionally, further analysis found a stronger association in patients with both CFHR1 deletion and anti-FH autoantibodies in comparison with only CFHR1 deficiency, indicating that CFHR1 deficiency might integrate with anti-FH autoantibodies, which contributes to the development of aHUS.The results from this comprehensive meta-analysis demonstrated that the deletion of CFHRs do encode proteins (FHR1-5) that are likely to modify the function of CFH and regulate complement activity, which would maintain the balance of the complement system and prevent humans from damaging aHUS. In concordance with previous studies that reported CFHR1 deletion was relevant to aHUS susceptibility, the results of present meta-analysis showed a significant association between CFHR1 deficiency and an increased risk of aHUS. The potential mechanism of FHR1 on complement activation and regulation might be via its ability to mediate CFH activity [CFHR1 deficiency may induce auto-antigenic neoepitope which is related to the formation of autoantibodies, and reduces the function of CFH. Meanwhile, subtotal CFH insufficiency can give rise to aHUS after a complement-activating trigger within the kidney and that the latter is C5 dependent [CFHRs and CFH have several independent large genomic duplications, such repeats tend to occur in nonallelic homologous recombinations, which results in the loss of CFHR1 [CFHR1 deficiency and the increased risk of aHUS, the present studies showed that individuals carrying CFHR1/R3 deficiency (without autoantibodies) had no significant association with aHUS. It is possibly influenced by the small number of included studies. Thus, further multi-centric studies investigating the relationship between aHUS and different combinations of CFHRs deficiency need to be conducted before definitive conclusions can be made.Insufficient inhibition of complement is associated with a quantity of autoimmune and inflammatory diseases’ pathological processes such as aHUS ,24. Seveactivity . Under nactivity ,30,31, wependent ,33. Moreependent . Cell suependent ,33. Moreof CFHR1 ,35. FHR1of CFHR1 ,35. In aof CFHR1 . In contCFHR1 deficiency were significantly associated with the increased risk of aHUS, which suggests that this autoantibody might cross-react with CFHR1 deficiency to promote the progress of aHUS. The underlying mechanism for such an observed relationship might be that such a deficiency may favor development of these specific autoantibodies. CFH and FHR1 have high degrees of sequence duplications, especially in SCRs 19–20 and SCRs 4–5 (autoantibody epitope site) [Evidence from a previous study found that autoantibodies to CFH were connected with the development of aHUS . The epipe site) ,38. Suchpe site) ,40.CFHRs deficiency and/or autoantibodies potentially has important clinical implications for investigating these risk factors in the early stage of aHUS and improving prognosis. Though the result from Holmes et al. suggests that CFHR1/R3 deletion in the normal population is different worldwide. Depending on ethnic origin, the highest frequency for the deletion (54.7%) was seen in Nigeria [CFHRs deficiency and autoantibodies could also be considered as a surrogate biomarker for some clinically relevant incidents or aHUS susceptibility, especially in designing new diagnostic approaches, identifying susceptible people, and monitoring the treatment effects.Findings from this meta-analysis have important public health significance. Cases of aHUS generally have a poor outcome ,41. Ther Nigeria . CFHRs dCFHRs deficiency, anti-FH autoantibodies, and aHUS adjusted for these covariates. Thereby, other complement factors, genetic abnormalities, or other confounding factors might affect the results of the present analysis. Nevertheless, the qualities of most of the eligible studies were high, which adds to the strength of our analysis. Third, the present results were also likely to be affected by different separate examination methods of CFHR1. The way to examine the CFHR1 status of the patients and controls, varies from one study to another, and that might affect the present results as a potential confounding factor. Furthermore, one study used western blot to identify CFHR1 deficiency. As aHUS could be treated by administration of blood derived products, the approach by western blotting might be at high risk of false results in aHUS. Although this study should not substantially affect the overall qualitative inference, we cannot exclude the potential influence which might have led to underestimation or overestimation of the association. Fourth, the pooled results were mainly conducted in Western countries, which limited the generalization of findings. Last, although statistical tests did not suggest the presence of publication bias for the present study, we could not exclude all publication bias.Several potential limitations should be also considered in interpreting the results of this meta-analysis. First, as no available cohort studies were found, this meta-analysis only extracted data from case-control studies. Retrospective study is subjected to internal methodological deficiencies, which may limit the power of this meta-analysis. Second, our results were based on unadjusted estimates, potential confounding components such as pregnancy, family history, drugs, other complement factor genes, and other genetic abnormalities  likely aCFHR1 deletion, anti-FH autoantibodies and aHUS. CFHR1 deficiency was significantly associated with increased risk of aHUS. Moreover, this study also identified potential interactions between anti-FH autoantibodies and CFHR1 deletion in aHUS patients, indicating that the development of aHUS could be exacerbated via interaction between CFHR1 deficiency and anti-FH autoantibodies. The present meta-analysis provides evidence of the association between


PSA) gene -158G/A polymorphism with benign prostatic hyperplasia (BPH) susceptibility need a meta-analysis to obtain a more accurate conclusion. A systematic search was conducted in electronic databases for the collection of eligible studies on PSA -158G/A polymorphism and BPH susceptibility. Pooled odds ratios (ORs) and 95% confidence intervals (95% CIs) were then calculated. 7 case-control studies with 758 cases and 752 controls were included into the present meta-analysis. The analysis results showed no significant relationship between PSA -158G/A polymorphism and BPH susceptibility in total analysis. Interestingly, after subgroup analyses based on ethnicity and source of control, the polymorphism reduced the susceptibility of BPH in Caucasian group , but it increased the disease susceptibility in Asian  and population-based  groups. PSA-158G/A polymorphism may be an inhibitor to the incidence of BPH in Caucasians, but it is likely to be a susceptible factor in Asians.Inconsistency between reported findings on the association of prostate specific antigen ( Prostate glands are the largest accessory glands in male genital organs, with pivotal physiological functions. Benign prostatic hyperplasia (BPH) is the most common disease causing urination disorder in medium-elderly males, and its morbidity is increasing with age –2. BPH uPSA is located on the long arm of chromosome 19, and contains 5 exons, 4 introns and 3 promoters. Documents have demonstrated that the polymorphism-158G/A in this gene can potentially affect the transcriptional control of androgen over expression, and thus exert certain influences on PSA expression and prostate development [Prostate specific antigen (PSA) is a glucoproteinase, and its active form is a serine protease composed by 237 residues . 70% of elopment –14.PSA gene -158G/A polymorphism on BPH risk, but the findings remained contradictory. Consequently, present meta-analysis was carried out backed by all available studies to systematically clarify such relationship.Association researches have previously discussed the effect of n = 1), review (n = 1) and without controls (n = 13). Therefore, a total of 7 independent studies [As presented in Figure  studies –21 were  studies .PSA -158G/A polymorphism didn't affect the susceptibility of BPH significantly under all genetic models in total analysis. However, after subgroup analyses by ethnicity and source of control, PSA -158G/A polymorphism was correlated with decreased susceptibility of BPH in Caucasian group [AA vs. GG: OR = 0.47, 95% CI = 0.25-0.89 , AA+GA vs. GG model (P = 0.002), allele A vs. allele G model (P = 0.002) and GA vs. GG model (P = 0.005). Thus, the random-effects model was selected to evaluate overall results in these four comparisons, while the fixed-effects model was used for the other model (P = 0.314). As for the significant heterogeneity, it was dramatically attenuated or completely removed after stratification analyses by ethnicity and control source. We therefore hypothesized that its plausible origins might include these two aspects.In Q test, significant heterogeneity was detected in AA We omitted individual studies one at a time and re-calculated overall estimates to detect the impact of each included study on pooled results. Since pooled ORs were not substantially altered, our results were statistically stable and credible.P = 0.150), indicating publication bias was negligible.Both Begg's funnel plot and Egger's test were utilized to assess the possibility of existing significant publication bias. All funnel plots displayed symmetrical shapes Figure , and EggBPH is a chronic progression disease, involving multiple factors and bothering elderly men in general –23. MultPSA gene is polymorphic, and among others, a polymorphism at position -158 (G/A) is one of the most frequently studied objects. In a Turkey population, Gunes et al. found an increasing effect of the GG genotype of PSA -158G/A polymorphism on BPH susceptibility, with an age-adjusted OR of 2.77 (95% CI = 1.277-6.012) [PSA -158G/A polymorphism between controls and BPH cases in Turks, suggesting that the polymorphism might not have independent influence on the disease onset [PSA, regulated by androgen, is produced by prostate cells, and is generally used as a prostate disease marker to predict the enlargement of prostate volume clinically. 7-6.012) . On the 7-6.012) . Whereasse onset .PSA -158G/A polymorphism and BPH susceptibility for a more convincing conclusion in view of distinct discrepancies among those findings. In our study, PSA -158G/A polymorphism showed no significant impact on BPH incidence in total analysis, but we observed with great interest that this polymorphism exerted opposite effects in different subgroups after stratification analysis by ethnicity and source of control. Specifically, under the genetic comparisons of AA vs. GG and A vs. G, PSA -158G/A polymorphism reduced the disease risk in Caucasian group, and increased the risk in Asian and population-based groups. To verify the reliability of the overall estimates from the present meta-analysis, we also implemented a series of tests. In sensitivity analysis, no qualitative alteration in pooled ORs was observed after the omission of any one of included studies, confirming the robustness of final outcomes. Besides, both funnel plot and Egger's test offered evidences supporting the absence of significant publication bias among selected studies. Hence, our study results had certain powerful in themselves.It is necessary to perform a meta-analysis so as to combine those conflicting findings on the relationship between Nevertheless, several limitations still needed to be acknowledged. First of all, we only included the articles previously published through online database searching, so possibly relevant reports unpublished or in other sources might be missed. Secondly, only seven eligible studies were ultimately incorporated into the current meta-analysis, with a relatively small sample size, and this situation might affect the precision of our results. Thirdly, not all eligible provided data adjusted for potentially confounding factors, which might introduce some bias into the final results. Fourthly, effects of gene-environment and gene-gene interactions on BPH susceptibility were not considered in this paper owing to inadequate data. Finally, since meta-analysis is a secondary analysis, the overall estimates may be affected by the quality of included studies. Therefore, our findings should be interpreted prudently.PSA -158G/A polymorphism may be a protecting factor against BPH in Caucasian populations, but it may enhance the disease risk in Asians. In view of the shortcomings mentioned above, well-designed studies with larger sample sizes are warranted to verify these findings in future.In short, this meta-analysis suggests that PSA gene -158G/A polymorphism with BPH susceptibility; (ii) case-control studies; (iii) sufficient data for computing crude odds ratios (ORs) with their 95% confidence intervals (95% CIs). Papers not satisfying any one of the above standards were excluded from the current study. In addition, removed publications also contained letters and reviews.Studies were included into this meta-analysis if they met the following criteria: (i) exploring the correlation of Relevant studies were searched from PubMed, EMBASE, Chinese National Knowledge Infrastructure (CNKI) and Web of Science databases. Combinations of the following terms were adopted: “prostate specific antigen” or “PSA” or “KLK3” or “-158G/A”, “polymorphism” or “mutation” or “variant”, and “benign prostatic hyperplasia” or “BPH” or “benign prostate hyperplasia”. References cited in pertinent articles were also screened manually to search for additional relevant papers.P values for Hardy-Weinberg equilibrium (HWE) in controls. All discrepancies over extracted data were resolved via discussion between the two reviewers. If not, the third reviewer would be consulted to reach a final determination.Two reviewers abstracted the following information independently from included studies: first author's name, year of publication, original country, ethnicity, control source, genotyping method, numbers of cases and controls, genotype frequencies in case and control groups, and P > 0.05 were regarded as fine compliance to the Law. Pooled ORs with 95% CIs were calculated to estimate the relationship between PSA -158G/A polymorphism and BPH susceptibility under five models: AA vs. GG, AA+GA vs. GG, AA vs. GG+GA, allele A vs. allele G and GA vs. GG. The significance of the pooled ORs was determined by Z-test. Subgroup analyses according to ethnicity and source of control were performed so as to further test such relationship. Heterogeneity among the included studies was checked by chi-square based Q-test, and P > 0.05 in the test suggesting the absence of significant heterogeneity among studies determined that the pooled ORs were computed with fixed-effects model [P < 0.05) [All the statistical tests were conducted using the STATA  software. The conformity of genotype distribution to HWE in controls was investigated through the goodness of fit chi-square test, and ts model . Otherwi < 0.05) . Sensiti < 0.05)  were use


The KIF6 rs20455 polymorphism has been verified as an important genetic factor of coronary heart disease (CHD), but with controversial results. The aim of this study was to explore the association between KIF6 rs20455 polymorphism and susceptibility to CHD.All eligible studies were identified by searching Medline (mainly PubMed), EMBASE, the Web of Science, Cochrane Collaboration Database, Chinese National Knowledge Infrastructure, Wanfang Database and China Biological Medicine up to October 5, 2016.Odds ratios (ORs) with 95% confidence interval (CI) were used to explore the association between KIF6 rs20455 polymorphism and CHD risk. Begg’s and Egger’s tests were used to examine the publication bias. Subgroup analysis and sensitivity analysis were performed to test the reliability and stability of the results. All the analyses were carried out by Stata 12.0 software.P = 0.801; Heterozygote model: OR = 1.009, 95% CI = 0.968–1.052, P = 0.636; Dominant model: OR = 1.007, 95% CI = 0.966–1.048, P = 0.753; Recessive model: OR = 0.989, 95% CI = 0.943–1.037, P = 0.655; Allele comparison model: OR = 1.00, 95% CI = 0.971–1.030, P = 0.988). Furthermore, subgroup analyses were performed by ethnicity, source of control.A total of 28 publications including 50 individual studies were analyzed in this present work. There are no significant association found between KIF6 rs20455 polymorphism and CHD risk (Homozygote model: OR = 1.007, 95% CI =0.952–1.066, Our result suggests that KIF6 rs20455 polymorphism may not be associated with CHD susceptibility. However, additional very well-designed large-scale studies are warranted to confirm our results. Coronary heart disease (CHD), a multifactorial heart disorder resulting from both environmental and genetic factors , is one The KIF6 protein is one of several molecular components that mediate intracellular transport of organelles, protein complexes, and mRNAs. A common Trp719Arg (rs20455) SNP in exon 19 of the KIF6 gene has been identified as a potential risk factor for CHD . The KIFA comprehensive search for all related studies from both electric databases, such as, Medline (mainly PubMed), Embase, Web of science, China National Knowledge Infrastructure (CNKI) et al., and hand search from references of all eligible literatures. Single or combinations of the following keywords were used: “kinesin like protein 6” or “KIF6” or “rs20455” or “719Arg”, “single nucleotide polymorphism, SNP or variation, mutation”, “genetic association” and “coronary heart disease” or “CHD”. No language and sample size were set. When more than one studies of the same population were included in several publications, only the most recent or complete studies were included in this meta-analysis.Articles included should meet following criteria: an appropriate description of KIF6 rs20455 polymorphism in CHD cases and healthy controls; results expressed as odds ratio (OR); and studies with a 95% confidence interval (CI) for OR with sufficient data to calculate these numbers. While for the exclusion criteria provided as follows: studies without raw data; case-only studies, family-based studies, case reports, editorials, and review articles . In studies with overlapping cases/controls, the study with the higher quality score or the study with more information on the origin of the cases/controls was included in the meta-analysis.Two researchers extracted important information independently and carefully from all eligible studies according to the criteria listed above. Any disagreement will be resolved by the two authors through discussion or the third author. The following data were extracted from each included study: first author’s surname, year of publication, country, ethnicity, genotyping method, source of control, total number of cases and controls, distributions of KIF6 rs20455 genotypes. Different ethnicity descents were categorized as Caucasian, Asian, and Mixed populations .P value was <0.05 will be considered statistically significant. Heterogeneity was also assessed using the I2 statistic, which takes values between 0% and 100% with higher values denoting greater degree of heterogeneity  [P value of heterogeneity was <0.05, or the I2 > 50%. Otherwise, fixed effects model (Mantel-Haenszel method) will be adopted [P value of <0.05 was considered significant. Subgroup analyses were performed by ethnicity, source of control, to confirm if our results were stable and robust [P values were two-sided.We adopted poled ORs and corresponding 95% confidence interval (CIs) to detect the association between KIF6 rs20455 polymorphism and CHD risk. Heterogeneity was explored by Q statistic , and thegeneity) . Differegeneity)  will be  adopted . Five ged robust . Begg’s d robust  and Egged robust  were expd robust . All theTotally, 209 potential relevant studies were searched through several databases. Based on the including criteria listed above, only 28 articles including 50 separate studies were included finally . A flow I2 = 33.9%; Pheterogeneity = 0.012), heterozygote , dominant , recessive  and allele comparison model  between this gene variation and the risk of CHD. So, random-effect model was used to calculate the statistical parameters. Overall, there were no significant association existed between KIF6 rs20455 polymorphism and the risk of CHD  and Egger’s regression . All the results are listed in Table There was no significant publication bias found in the meta-analysis, reflected by Large sample and unbiased epidemiological studies of predisposition genes polymorphisms could provide insight into the in vivo relationship between candidate genes and complex diseases. Many epidemiological studies have investigated the relationship between the KIF6 rs20455 polymorphism and the risk of CHD, but because of small sample size and the low statistical power of individual studies, results have been contradictory. In this present study, we searched all eligible studies to date and got the precise result if KIF6 rs20455 polymorphism could contribute to the risk of CHD. To the best of our knowledge, our present work was the most comprehensive one through enrolling all eligible studies.Herein, we included 50 individual studies, including 40,059 cases and 64,032 controls. Overall, there was no association between KIF6 rs20455 polymorphism and CHD risk. Hamidizadeh et al. found that significant association was found between this gene polymorphism and CHD risk among Caucasian populations , and thePublication bias was found in some genetic models. The explanations might arise from some aspects. First, our meta-analysis took into consideration only fully published data, and the abstract and conference papers were excluded. Second, this meta-analysis only focused on papers published in Chinese and English language, and some eligible studies which were reported in other languages might be missed. Third, positive results tend to be accepted by journals while negative results are often rejected or not even submitted. We should point out that the publication bias might partly account for the results, but which were not affected deeply. When we adjusted the results using the trim and fill method, the adjusted risk estimate was attenuated but remained significant, indicating the stability of our results.Some limitations of this meta-analysis should be addressed. Firstly, heterogeneity is a potential problem when interpreting all the results of meta-analysis. Although we minimized the likelihood by performing a careful search for published studies, using the explicit criteria for study inclusion, the significant between-study heterogeneity still existed in most of comparison. The presence of heterogeneity can result from differences in the age distribution, selection of controls, prevalence lifestyle factors and so on. Secondly, only published studies were included in this meta-analysis. Therefore, potential publication bias was existed in some genetic models. Despite the limitations, our meta-analysis significantly increased the statistical power based on substantial data from different studies. The sensitivity analyses outcomes reflected that our results were statistically stable and reliable.In conclusion, this present meta-analysis suggests that carriers of KIF6 rs20455 polymorphism may irrelative to the risk of CHD. We also observed no compelling evidence of an association between the KIF6 rs20455 SNP and CHD in multiple race/ethnic groups. These findings do not support the clinical utility of testing for the KIF6 rs20455 polymorphism in the primary prevention of CHD and indirectly question whether genotype information at this locus is able to identify subjects most likely to benefit from the use of statins.


Various literatures have demonstrated that overexpression of Metadherin (MTDH) is correlated with tumor metastasis and it can predict poor survival outcomes in female reproduction malignancies. In order to enhance the statistical power and reach a recognized conclusion, we conducted a systematic review and meta-analysis to thoroughly investigate the association of MTDH expression with tumor metastasis and survival outcomes following PRISMA guidelines. Odds ratios (ORs) and hazard ratios (HRs) were used to demonstrate the impact of MTDH on tumor metastasis and prognosis respectively. Data were pooled with appropriate effects model on STATA12.0. Our results indicated that high MTDH expression is significantly correlated with higher mortality for breast, ovarian and cervical cancer. High immunohistochemical expression of MTDH is remarkably associated with shorter disease-free survival (DFS) in breast cancer but not in ovarian cancer. The pooled results suggested that high level of MTDH significantly predicted distant metastasis and lymph node metastasis in breast cancer. Strong associations were observed between MTDH expression and lymph node metastasis in ovarian and cervical cancer. In conclusion, MTDH might be a novel biomarker which can effectively reflect metastasis status and prognosis of breast cancer. However, its application in clinical practice needs more prospective studies with large samples. Reproduction malignancies, including breast cancer, cervical cancer, ovarian cancer and endometrial cancer, have been one of the major causes of death in females, among which, breast cancer is the most common malignancy and the leading cause of cancer death in western countries. With an increasing incidence, reproduction malignancies have seriously affected living quality and health of the patients. It is estimated that there were 1.7 million new cases, causing 521,900 deaths throughout the world in 2012Recently, researchers have made great efforts to explore new biomarkers which are associated with the diagnosis, progression and prognosis of reproduction malignancies. Metadherin (MTDH), as a cell surface protein, could induce breast cancer cells transferring to lung in mouse model. Its gene, astrocyte elevated gene-1 (AEG-1), was first cloned in human fetal astrocytes as an inducible gene by human immunodeficiency virus 1 (HIV)-1 and tumor necrosis factor-α (TNF-α) in 200239Up till now, no quantitative evaluation was performed. Because of the limited sample size, the conclusion of a single study lacks power of reliability. Hence, by reviewing published literature we performed a comprehensive meta-analysis in order to get a consistent and reliable conclusion and to cast light on the impact of MTDH expression on metastasis and survival status.In total, 1115 studies (791 in English and 324 in Chinese) were retrieved for our systematic review and subsequent meta-analysis after searching different databases. After removing duplicates, titles and abstracts of the remained 765 papers were prudently screened, among which 74 studies were obtained as they evaluated the role of MTDH/LYRIC/AEG-1/3D3 in female reproduction malignancies. Meanwhile, a total of 55 potentially relevant papers were excluded as they were against the inclusion criteria of our meta-analysis, of which three studies only investigated the gene expression of MTDH in breast and ovarian cancer. Two studies1256781415161718192021222324252627The 19 studies included in our meta-analysis analyzed the correlation of MDTH levels with clinicopathological parameters and survival outcomes in 2483 female reproduction malignancies patients. Among them 8 studies were about breast cancer514151621232561718202226719A large proportion of these trails were performed in China and the number of patients ranged from 60 to 225. As for inclusion criteria, all the samples of included studies were tissues and their MTDH expression was examined by immunohistochemical (IHC). Eleven studies568151617182123257141920222427Data of distant metastasis were available in 8 studies51415161719232557815161719232526275678141617182023242526681620242527P < 0.001) , the fixed effects model was applied. Four studies162325P < 0.001) . Seven studies51416192325P < 0.001) . Six studies515162325I2 = 38.8%; P = 0.147) and the pooled fixed effects HR was 1.880  (Eight studies51415162123255141623< 0.001) , Table 3< 0.001) , Table 3< 0.001) , Table 3< 0.001) , Table 3I2 = 92.5%; P < 0.001), and the pooled HR was 4.525 (P = 0.060)  for these 3 studies. There were another 3 studies1726P = 0.002) , we applied the random-effects model.Seven studies61718202226617202226< 0.001) , Table 3= 0.060) , Table 3= 0.002) , Table 3P < 0.001) , thus we performed the random-effects model. To examine the relationship between high tumoral MTDH expression and lymph node metastasis in cervical cancer, we combined OR from two studies7P = 0.001) .Three studies7197< 0.001) , Table 3= 0.001) , Table 3P < 0.001) . A total of 8 studies681620232527P < 0.001) . The pooled results from these studies indicated MTDH overexpression in reproduction cancers can cause a short lifespan for patients. Besides, 8 studies51416171923255781516171923252627P < 0.001) (P < 0.001) , while moderate heterogeneity can be found in the studies related to lymph node metastasis .Based on 14 studies that examined the association of MTDH with mortality of various female reproduction cancers, the pooled random effects HR was 3.647  , Table 3< 0.001) , Table 3< 0.001) , Table 3et al. significantly affected the pooled HR  to 3.300  (P < 0.001).After excluding every single study in order, the pooled HRs for mortality and ORs metastasis did not change significantly . For DFSooled HR . After e< 0.001)  and no hP = 0.112). However, significant publication bias was observed for the studies concerned with lymph node metastasis (P = 0.001). Trim and fill method was conducted to enroll missing studies. After enrolling missing studies, the pooled random-effects OR was 2.150 (95% CI: 1.465–3.155)  . Due to et al. showed that high MTDH expression was remarkably associated with short DFS in breast cancer (P = 0.0233) with a cohort of patients from public data (GSE1378). The study published by Liu Pet al. suggested that high MTDH level was significantly correlated with poor OS in triple negative breast cancer (P = 0.006), but no statistical significance was observed for DFS (P = 0.065).Two studies12Meta-analysis can enhance statistical power by quantitatively combining the results of multiple independent studies. As several studies have regarded MTDH as the potential biomarker which can indicate the metastasis and prognosis in malignancies, and as MTDH could miraculously induce breast cancer cells transferring to lung in mouse model, we carried out this comprehensive systematic review and meta-analysis based on published literature to thoroughly investigate the significance of MTDH in reproduction malignancies. Our results showed that MTDH, as a cell surface protein, is significantly associated with the mortality of patients with reproduction malignancies (HR = 3.647), including breast cancer HR = 2.728), ovarian cancer HR = 4.525), cervical cancer (HR = 2.524) and endometrial cancer (HR = 2.524). For DFS, significant result was observed in breast cancer (HR = 2.966) but not in ovarian cancer. Moreover, high MTDH expression remarkably increased the risk of distant metastasis (HR = 3.739) and lymph node metastasis (HR = 2.696) in reproduction malignancies. A strong association was observed between MTDH and metastasis as well as prognosis in clinical malignancies. The pooled results of those 8 studies25, cervi141516212325, ovarianI2 = 82.1%) and DFS (I2 = 80.7%). However, when we divided the included studies into different cancer types, no significant heterogeneity was found in breast cancer (P > 0.05). Thus, the difference in cancer type might be one of the potential factors that caused inter-study heterogeneity. For ovarian cancer, the possible reasons of heterogeneity might be the selection of tissue and the computing method of HR. Of 7 studies in ovarian cancer, 3 studies617182022et al. only analyzed the prognosis of patients with stages III–IV. The study published by Dong Ret al. only provided survival curve while the other studies offered HR value of multivariate analysis. To take all the heterogeneity described above into consideration, we pooled the HRs using random effects model. For the combination of those studies which didn’t have statistically significant heterogeneity, a fixed effects model was used to acquire more accurate combined HR.Heterogeneity is one of possible factors impacting the pooled results of meta-analysisP = 0.112) was observed in mortality group while significant publication bias was observed in lymph node metastasis group (P = 0.001). Thus, trim and fill method was conducted to enroll missing studies. However, the final result was not distinctly altered, which implied that the pooled result was stable and creditable. The results of sensitivity analysis indicated that our meta-analysis was stable for all the groups except DFS. For DFS, the result didn’t change obviously after eliminating the study published by Dong Ret al. Meanwhile, the result of sensitivity analysis for studies using the same cut-off (SI ≥ 4) was similar to all studies of mortality group, which implied the reliability of our conclusion.The unpublished studies or studies with negative results were not included in our meta-analysis, which might cause same bias and affect the pooled results. For publication bias, we performed Begg’s test in the mortality and lymph node metastasis group. No obvious publication bias , which favorably avoided the effects of small sample. Thirdly, all the samples of the included studies were from tissues and their MTDH was examined by IHC, which ensured the homogeneity of our meta-analysis. Moreover, there was no significant heterogeneity when we pooled the studies in breast cancer. The strengths above immensely enhanced the reliability of our meta-analysis. Still, there were also a few limitations. Four studies232425et al. For reproduction malignancies, it was reported that inhibition of MTDH increases cancer cells’ sensitivity to chemotherapy drugs, such as AZD6244, tumor necrosis factor-a-related apoptosis-inducing ligand (TRAIL) and HDAC inhibitor LBH5893132et al. confirmed that MTDH induces radioresistance and inhibits apoptosis in cervical cancer cells3435363738364121Recently, a number of studies have investigated the impact of MTDH in reproduction malignancies. They rifely concluded that MTDH was associated with specific biological characteristics and molecular pathways related to tumor deterioration, but the exact mechanism was still unclear. Chemoresistance/radioresistance might be one of the reasons related to poor prognosis. MTDH could mediate drug resistance in various cancersIn conclusion, our meta-analysis suggests that high MTDH expression increases risk of distant metastasis and lymph node metastasis in reproduction malignancies. Furthermore, high MTDH expression is remarkably associated with high mortality in reproduction malignancies. Specifically, the results our meta-analysis indicates that high MTDH expression is strongly associated with high risk of tumor metastasis and worse prognosis in breast cancer. Thus, our study provides evidence that MTDH might be a potential novel biomarker effectively reflecting metastasis status and prognosis of breast cancer patients, which might help to formulate a better therapy for individual patient. Simultaneously, to enhance the creditability of our meta-analysis, more prospective studies with large samples are needed.A comprehensive electronic literature search was conducted using the following English databases and Chinese databases: PubMed, Wiley Online Library, Science Direct, EMBASE, Cochrane Central Register of Controlled Trials, ISI Web of Science, China National Knowledge Infrastructure (CNKI) and Wanfang database. The literature search strategy was as follow: “metadherin OR MTDH OR astrocyte elevated gene-1 OR AEG1 OR AEG-1 OR LYRICOR LYRIC/3D3 OR 3D3″[All Fields] AND “tumor OR tumour OR cancer OR carcinoma OR neoplasm OR neoplastic OR malignancy”[All Fields] AND “prognosis OR prognostic OR survival OR survivance OR mortality OR outcome OR predict* OR follow-up” [All Fields]. The search was not finished until September 21, 2016.Firstly, all the titles and abstracts of each literature found by the search strategy were imported to EndNote X7. Then, two independent authors (Yongbin Hou and Liyi Hu) screened out the eligible literature which investigated the association between MTDH/LYRIC/AEG-1/3D3 expression and clinicopathological features as well as survival outcomes in patients with female reproduction cancers. Finally, full-text articles of eligible literatures were reviewed in detail, including references cited in the literature, according to our inclusion and exclusion criteria.The detailed inclusion criteria were listed in The following information was extracted from each study by two independent authors (Yonghua Mi and Yongbin Hou) using a standard excel form: first author; year of publication; types of cancer; source and number of patients; detected method for MTDH/LYRIC/AEG-1/3D3; cut-off value for positive MTDH/LYRIC/AEG-1/3D3 expression; longest follow-up time. For metastasis status of included studies, ORs as well as their 95% CI were extracted. For survival analysis, HRs along with their 95% CI and P value were extracted. If the results multivariate analysis were available, the HRs/ORs values of multivariate analysis and concomitant variables were extracted. Otherwise, the HRs/ORs of univariate analysis  were extracted. Subsequently, all the data were double-checked by a third author (Liyi Hu) and controversies were solved by discussion.For the meta-analysis, we stratified the results by the type of reproduction carcinoma. To describe the intensity of relationship between MTDH and clinical outcomes in reproduction malignancies patients, HR value and OR value as well as their 95% CI were used. With a standard excel form, the HR value and OR value of each single study were extracted. For HR and OR value, we extracted the relevant data directly if they were given in the studies. As for those statistical variables which were indirectly stated, we estimated their values on the basis of available data or Kaplan-Meier curves by methods of Jayne F TierneyI2 statistical methodP < 0.05), the random-effects modelP > 0.05). The I2 statistic was regarded as a quantitative measure of the degree of inconsistency among the studies. When I2 = 0, there was no heterogeneity. The larger the percentage, the increasing heterogeneityI2 value of 1 to 25%, 25 to 75%, 75to 100% was considered low heterogeneity, moderate heterogeneity and high heterogeneity, respectively. A statistically significant HR (OR) > 1 suggested that high MTDH indicated worse survival outcomes and increased the risk of metastasis. P values less than 0.05 were considered to be statistically significant.Subsequently, the HRs (ORs) values were quantificationally combined on STATA12.0 . For heterogeneity analysis, the Cochran Q test and the Higgins Concerning the positive results being more likely to published, publication bias was assessed by Begg’s test with funnel plots when the meta-analysis included 10 or more studies. A symmetrical inverted funnel implied that there was no significant publication bias among included studies. Trim and fill method was conducted to reduce publication bias if statistically significant publication bias was observed. Furthermore, sensitivity analysis was conducted by eliminating single study successively and the rest studies was pooled to examine the stability of the results. Also, sensitivity analysis was performed in studies of mortality which employed the same cut-off value (SI ≥ 4). The current study is a meta-analysis which gained ethical approval and all the procedures of this study followed PRISMA guidelinesHow to cite this article: Hou, Y. et al. Association of MTDH immunohistochemical expression with metastasis and prognosis in female reproduction malignancies: a systematic review and meta-analysis. Sci. Rep.6, 38365; doi: 10.1038/srep38365 (2016).Publisher's note: Springer Nature remains neutral with regard to jurisdictional claims in published maps and institutional affiliations.