The air pollution emitted by petrochemical industrial complexes (PICs) may affect the respiratory health of surrounding residents. Previous meta-analyses have indicated a higher risk of lung cancer mortality and incidence among residents near a PIC. Therefore, in this study, a meta-analysis was conducted to estimate the degree to which PIC exposure increases the risk of the development of nonmalignant respiratory symptoms among residents. We followed the Preferred Reporting Items for Systematic Reviews and Meta-Analyses (PRISMA) guidelines to systematically identify, select, and critically appraise relevant research. Finally, we identified 16 study groups reporting 5 types of respiratory symptoms: asthma, bronchitis, cough, rhinitis, and wheezing. We estimated pooled odds ratios (ORs) using random-effect models and investigated the robustness of pooled estimates in subgroup analyses by location, observation period, and age group. We determined that residential exposure to a PIC was significantly associated with a higher incidence of cough (OR = 1.35), wheezing (OR = 1.28), bronchitis (OR = 1.26), rhinitis (OR = 1.17), and asthma (OR = 1.15), although the latter two associations did not reach statistical significance. Subgroup analyses suggested that the association remained robust across different groups for cough and bronchitis. We identified high heterogeneity for asthma, rhinitis, and wheezing, which could be due to higher ORs in South America. Our meta-analysis indicates that residential exposure to a PIC is associated with an increased risk of nonmalignant respiratory symptoms. The petrochemical industry is frequently a large contributor to a country’s economic development and is a source of employment opportunities for communities . HoweverResidential exposure to PICs is significantly associated with various respiratory symptoms ,12,13,14By contrast, studies elsewhere have not demonstrated significant associations between residential proximity to a PIC and the aforementioned respiratory symptoms ,16. A stTwo meta-analyses investigated the association between PIC exposure and lung cancer mortality and incidence, respectively ,18. HoweWe followed the Preferred Reporting Items for Systematic Reviews and Meta-Analyses (PRISMA) guidelines to search literature and select studies. We defined “ AND ((respiratory symptom) OR )” as our search terms. Two authors (W.W.C. and H.B.) separately searched for studies available in the period January 1, 1971 to January 9, 2020 in PubMed, Web of Science, EBSCOhost, Cochrane Library, ClinicalKey, and Embase databases.These studies reported eight types of nonmalignant respiratory symptoms. After the exclusion of three types of symptoms within ≤2 study groups ,14, our 2 test was applied to address heterogeneity between the collected study groups, with 10% indicating no heterogeneity, 10–30% indicating low heterogeneity, 30–60% indicating moderate heterogeneity, and >60% indicating high heterogeneity [We obtained prevalence, adjusted odds ratios (ORs), and 95% confidence intervals (CIs) for the 16 study groups. We organized data according to the five aforementioned respiratory symptoms and applied random-effect models to analyze the pooled risk of the respiratory symptoms in residents near PICs, which accounted for between-study variation of heterogeneity ,25. Thisogeneity . We usedogeneity .We performed subgroup analyses to examine the robustness of pooled estimates using different locations, observation periods, and age groups of the study populations. According to the median of observation periods reported in the 11 studies, we divided observation periods into two categories: ≥3 and <3 years. We reviewed population age in each study and defined the following age subgroups: 6–8 years representing the “junior group”, 11–14 years representing the “intermediate group”, and 0–14 years (or those whose ages are not known) representing the “wide group”. Finally, we applied a funnel plot to examine whether publication bias was present across studies.To assess the quality of the studies and incorporate quality assessment into our interpretation of the results, we used the Newcastle-Ottawa Quality Assessment Scale adapted for cross-sectional studies . We evalOur study was registered on PROSPERO on 26 September 2019. The registration number is 152438.2 was 0.0%, indicating no heterogeneity for cough in our meta-analysis. Due to its narrow 95% CI (at 1.00–1.82), the estimate of a study in Thailand accounted for the largest percentage weight (50.38%). The estimates for cough across the four study groups had no asymmetry  (= 0.868) A.2 was 71.0%, signifying high heterogeneity for wheezing in our meta-analysis. We observed that all Asian study populations had non-significant pooled estimates, whereas all study populations from the Americas had significant pooled estimates. Although the data collected from a Brazilian study had an OR of 2.00, the estimate for this study accounted for only 3.76% of the weight proportion because of its wide 95% CI (1.01–4.01). Publication bias was observed for wheezing among ten study groups because three were outside the confidence limits. Despite being outside the limits, two of the three study groups were from studies that had more than seven stars ; the other study group (from a satisfactory study) was closer to the confidence limits  (As presented in = 0.393) B. 2 was 13.1%, signifying low heterogeneity for bronchitis in our meta-analysis. All studies had pooled estimates greater than 1. The highest OR was observed for Taiwan at 1.99, but it accounted for only 7.63% of the proportion weight because of its wide 95% CI (at 1.21–3.28). The estimates for bronchitis among the six study groups had no asymmetry  (As displayed in = 0.054) C.2 was 88.5%, implying high heterogeneity for rhinitis in our meta-analysis. We observed that a European study population (aged 13–14 years) had pooled estimates that were less than 1. As detailed in p value = 0.479).As displayed in n = 6) had pooled estimates of less than 1, whereas the study populations of other regions (n = 4) had pooled estimates greater than 1. Furthermore, the study conducted in Taiwan had the highest OR (2.76), but it had the lowest weighting factor of 6.22% because of its wider 95% CI . The overall I2 was 82.4%, indicating high heterogeneity for asthma in our meta-analysis. As may be observed in p value = 0.767).As shown in As presented in As displayed in As presented in Environmental pollutants from PIC operations have been associated with an increased prevalence of respiratory symptoms ,12,14. B10 enter the respiratory tract, they may be stopped at the nasal cavity, thus causing rhinitis. Fine particles can enter further into the tracheobronchial tree (thus causing asthma), travel to the bronchioles and alveolar parts of the lung, and even enter the bloodstream [The order of the risk of the five respiratory symptoms approximated the order in which particles and chemical toxicants enter the respiratory tract and cause inflammation or other symptoms. Cough is a basic symptom of respiratory disease . When chodstream ,37. Pollodstream ,39. Thesodstream ,40. Thusodstream .Our subgroup analyses revealed that some groups bear a higher risk, namely those in studies conducted in South America or in studies with a longer observation period. This first result is comparable to the incidence rate estimated by the Global Burden of Diseases, Injuries, and Risk Factors Study . Taking However, our study has some limitations. First, the included articles compared risk based on areas or geographic regions as defined by the authors rather than on absolute distance to PICs. We may underestimate the risk of respiratory symptoms for people living closer to PICs. Future studies using distance as a continuous variable might provide a more accurate estimation. Second, potential confounders might not have been comprehensively considered and adjusted for in some of the original studies. For example, smoking was not adjusted for in all the studies. Although smoking is less prevalent among residents in counties with a refinery (54%) than in counties without one (70%) , we may The size of the global petrochemicals market is increasing. Our meta-analysis confirmed significant associations between residential proximity to PICs and the risk of nonmalignant respiratory symptoms. The order of the risk of five respiratory symptoms approximate the order in which particles and chemical toxicants enter the respiratory tract and cause inflammation or other symptoms. Our findings suggest additional in-depth studies are required to explore other reasons behind high respiratory risks among people in South America as well as to document and quantify the effect of each pollutant on impairments of respiratory health.


Intracranial aneurysm (IA) is one of the main causes of subarachnoid hemorrhage (SAH) leading to a high percentage of disability and mortality worldwide. In addition to environmental factors, the risk of rupture or prognosis of intracranial aneurysm is also closely related to gene. Thus, a lot of genetic studies have been used to explore associated risk genes as well as variant loci of intracranial aneurysm and found several chromosome variates including 9p21.3 (CDKN2BAS) related to Intracranial aneurysm. However, due to differences in population and the existence of SNP, it is still not determined that whether these genetic changes can be identified as independent risk factors for intracranial aneurysm. Therefore, we performed a meta-analysis of CDKN2BAS SNPs to explore its association with intracranial aneurysms and the results show a significance relation between rs10757272, rs1333040, and rs6475606 with intracranial aneurysm. This will open a new perspective for future intracranial aneurysm gene research and therapy. In addition to environmental factors including smoking, drinking, hypertension, diabetes, and so on, gene is also related to intracranial aneurysm. Since the family history of Intracranial aneurysm, genetics is considered as a major risk factor. The prevalence of first-degree relatives of SAH patients is 2 to 6 times higher than that of the control group.–8 Moreover, it also seems that a family history may be a crucial factor influencing the risk of aneurysm rupture or the prognosis of intracranial aneurysm. Thus, plenty of genetic studies have been performed to identify associated risk genes and loci of intracranial aneurysm during these years.Intracranial aneurysm (IA) are balloon-like dilations of the intracranial arterial wall that usually located at the branching points of the major arteries at the base of the brain.–12Genome-wide association study (GWAS) is a critical approach in genetic research. Several genes have been found to be closely related to intracranial aneurysms by using GWAS, including chromosome 4q31.23 (EDNRA), 8q12.1 (SOX17), 9p21.3 (CDKN2A/CDKN2B/CDKN2BAS), 10q24.32 (CNNM2), 12q22, 13q13.1 (KL/STARD13), 18q11.2 (RBBP8), and 20p12.1.3. The effect of genetic variants associated with intracranial aneurysm varies among different populations because of the different linkage disequilibrium (LD) structures across populations and potential interaction between genetic variants and environment factors.However, whether some of these chromosomal genes can be identified as true risk factors for intracranial arteries remains to be discussed due to the influence of population stratification and confounding nongenetic factors.,3 Recently, the studies of GWAS in Asian and Caucasian population identified several CDKN2BAS single-nucleotide polymorphisms (SNPs) associated with intracranial aneurysm. Hence, a meta-analysis was designed to explore the relationship between CDKN2BAS SNPs and intracranial aneurysms.CDKN2BAS located in chromosome 9p21, and it is associated with a variety of human diseases, including glioma, prostate cancer, stomach cancer, pancreatic cancer, leukemia, colorectal cancer, lung cancer, diabetes, and aneurysm.2Preferred Reporting Items for Systematic Reviews and Meta-Analyses (PRISMA) guidelines. Since this study is a meta-analysis, ethical approval is not required.The meta-analysis was performed based on 2.1We searched the following databases: PubMed, Chinese Biomedical (CBM), web of science, and EMBASE (from inception through March 2019). The following search terms were performed to identify the studies which investigate the association of CDKN2BAS SNP and intracranial aneurysm: ([polymorphisms OR variant OR variation OR genotype] AND [CDKN2BAS or Anril]) AND aneurysms. In case of the omissions of the potential studies, additional studies were identified via searching the reference lists of included studies and review articles.2.2Inclusion criteria:1.Studies investigating the association of the CDKN2BAS polymorphisms with intracranial aneurysm risk;2.The odds ratio (OR) or relative risk (RR) with 95% confidence interval (CI) could be acquired or calculated from included studies.3.Patients in eligible studies diagnosed by computed tomography angiography (CTA), magnetic resonance angiography (MRA) or digital subtraction angiography (DSA) has at least one intracerebral aneurysm.2.31.Animal studies.2.Case report.3.Review articles.4.Studies investigating a population with an underlying connective tissue disorder or other genetic disorders associated with intracranial aneurysm.2.4Three investigators  independently screened the titles and abstracts of the texts. Then, the full-text was further assessed. YFX independently extracted data from included studies and supplementary materials. The data including: Author name, Year of publication, Sample size (case/control), Mean age, Race, Gender, OR with 95% CI, SNPs, Genotype distribution in cases and controls. Inconsistencies were discussed and resolved by senior researches (NW).2.5I2 statistic was calculated to assess the degree of heterogeneity among the studies, significant heterogeneity was defined as P < .05 and I2 > 50%. Subgroup analysis was used to further identify the factors influencing heterogeneity. Publication bias were evaluated through Begg's funnel plot analysis and Egger's tests (significant: P < .1).The ORs and 95% CIs under allelic comparison were calculated, respectively. The strengths of the associations between CDKN2BAS polymorphisms and intracranial aneurysm were estimated via the above-mentioned ORs with 95% CI. All data were analyzed by using the STATA 12, a random-effects model (DerSimonian–Laird) was used to calculate the pooled ORs because of the possible heterogeneity. The 33.1,3 rs1333040-three studies,,3,12 rs6475606-two studies, and rs10733376-one study.,16As shown in Figure 3.2,16 investigated the association between CDKN2BAS polymorphisms and intracranial aneurysm in Caucasian population, others investigated in Asian population. The genotype data and the OR values under allelic comparison (TT vs TC & CC) from each clinical study were shown inTable The detailed data of the included study characteristics were shown in Table 3.3. Notably, the I2 statistic showed significant heterogeneity among the included clinical studies .In light of previous studies, the frequency of T allele is strongly associated with the risk of intracranial aneurysm among the SNPs of CDKN2BAS, including rs10757272, rs1333040, rs6475606, and rs10733376. Hence, allelic comparison (TT/TC + CC) was used to detect the correlation between T allele and intracranial aneurysm risks. The results indicated that the CDKN2BAS SNPs were significantly associated with intracranial aneurysm P = .458) and Egger's test (P = .152) indicated no significant evidence of publication bias  which indicated the significant heterogeneity among the included clinical studies. In order to figure out the parameters that may affect the stability and reliability of pooled analysis, the subgroup analysis was performed to investigate the potential factors causing the heterogeneity.The aforementioned results showed that ,16 others investigated the association between CDKN2BAS polymorphisms and intracranial aneurysm in Asian population.,3,12 As shown in Figure P < .001; Caucasian: OR, 1.35; 95% CI, 1.26–1.44; P < .001). Moreover, we also found that the heterogeneity shown no statistical difference in subgroups . Since the I2 value was significantly decreased compared to which in pooled analysis . It indicated that the heterogeneity was significantly decreased after subgroup stratification. These outcomes indicated that race might be an important factor influencing heterogeneity.As we know, several studies have proven the racial difference in diseases occurrence. Hence, we first speculated that race in each study may be an important factor influencing heterogeneity. As mentioned above, two studies investigated the association in Caucasian,P < .001). The I2 statistic also show no significant heterogeneity . Likewise, similar results were shown in rs1333040 group  and rs6475606 group . In rs10733376 group, the P values and I2 values were not shown because of the insufficient quantity (Subjects ≤2). These results indicated that different CDKN2BAS SNPs might be another important factor influencing heterogeneity.In addition, because of the variant CDKN2BAS SNPs in the present research, studies were divided into four subgroups . The results indicated that rs10757272 was strongly associated with intracranial aneurysm  and platelet reactivity in the European population.,19 In a study involved with the asymptomatic community-based Korean populations, researchers demonstrated that rs10757272 related to severe coronary artery calcification (CAC) based on each age and sex. This proves that it might be a common genetic risk factor in some cardio-cerebral-vascular diseases. However, due to the differences of genes in different populations, it is necessary to consider these heterogeneities in the analysis of the importance of these results. Though belonging to the Asian population same as Koreans and Japanese, no association of the variant rs10757272 with sporadic intracranial aneurysm in Chinese was shown in Yunchang Chen et al's study, the reason is probably due to genetic heterogeneity in different ethnic populations and may also be relevant in a bias caused by variation in the prevalence of a positive family history between populations.In this study, we aim to explore the CDKN2BAS SNPs  genetic relationship with intracranial aneurysm, the results showed that rs10757272 was strongly associated with the disease. CDKN2BAS, a 3.8 kb long non-coding RNA (lncRNA) that located in chromosome 9p21, Genetic variants of rs1333040 can also be an important risk factor in multiple intracranial and extracranial diseases. For example, Sturiale et al demonstrated an association between rs1333040 in the chromosomal 9p21 locus and sporadic brain arteriovenous malformations (BAVMs). This result was further confirmed by experiments conducted by Bendjilali et al. They found that these genetic found in BAVMs were similar to that in aneurysms, the reason may be attributed to the similar vascular pathology mechanisms between them. It is been also reported that rs1333040 affects the progression of coronary atherosclerosis and the probability of undergoing subsequent coronary artery revascularization after a first myocardial infarction whereas results of cardiovascular death was not affected. It is worth noting that rs6475606, another SNP we found to be statistically significant, has also been shown to be associated with perioperative myocardial injury after coronary artery bypass graft surgery. These results also give us inspiration that such genetic variants will not only be identified as risk factors for intracranial aneurysms, but also more widely involved in many other cardiovascular diseases.In addition to rs10757272, we found that rs1333040 and rs6475606 were also associated with intracranial aneurysms, which is consistent with previous experimental results, though the effect of them was less significant than rs10757272. In the study by Nakaoka et al, rs1333040 was identified as an independent predictor of intracranial aneurysms among other SNPs in the 9p21 region.There are still several limitations in the present meta-analysis. First, the whole analysis was based on the study level instead of individual level. Secondly, the quantity of included studies was not sufficient which may lower the credibility of results. In addition, the present study demonstrated that the variant CDKN2BAS SNPs and race were likely to be the potential sources of heterogeneity. However, the meta-regression was unable to be conducted because of insufficient data. That may cause the omission of other potential factors. Moreover, PCR approaches, study designs, sample sizes, etc varied a lot in the present study, these were also the inevitable flaws of meta-analysis. Hence, the conclusion of the present study should be interpreted and adopted with caution.Conceptualization: Ting Yu, Hailong Jiang, Ning Wang.Data curation: Ting Yu, Hailong Jiang, Yunren Fan, Yunfeng Xu.Formal analysis: Hailong Jiang, Yunren Fan, Yunfeng Xu.Investigation: Hailong Jiang, Ning Wang.Methodology: Ting Yu, Hailong Jiang, Ning Wang.Supervision: Ning Wang.Validation: Yunren Fan.Visualization: Yunfeng Xu, Ning Wang.Writing – review & editing: Yunfeng Xu, Ning Wang.Writing – original draft: Ning Wang.


It is understudied whether the posed association of oral antibiotics with colorectal cancer (CRC) varies between antibiotic spectrums, colorectal continuum, and if a non-linear dose-dependent relationship is present.Three electronic databases and a trial platform were searched for all relevant studies, from inception until February 2020, without restrictions. Random-effects meta-analyses provided pooled effect-sizes (ES) with 95% confidence intervals (CI). Dose–response analyses modelling the relationship between number of days exposed to antibiotics and CRC risk were extended to non-linear multivariable random-effects models.Of 6483 identified publications ten were eligible, including 4.1 million individuals and over 73,550 CRC cases. The pooled CRC risk was increased among individuals who ever-used antibiotics , particularly for broad-spectrum antibiotics , but not for narrow-spectrum antibiotic . The dose–response analysis did not provide strong evidence of any particular dose–response association, and the risk patterns were rather similar for colon and rectal cancer.The antibiotic use associated CRC risk seemingly differs between broad- and narrow-spectrum antibiotics, and possibly within the colorectal continuum. It remains unclear whether this association is causal, requiring more mechanistic studies and further clarification of drug–microbiome interactions. Gut microbes have been associated with CRC promotion, particularly certain strains of Escherichia coli and Fusobacterium nucleatum. The latter might even involve a worse prognosis, and higher treatment resistance, if expressed at high levels.8Colorectal cancer (CRC) is the fourth most common cause of death globally,10 highlights the need for better understanding of the possible role of antibiotics and gut microbiome on the risk of CRC. Emerging evidence supports the hypothesis of oral antibiotics changing the gut microbiota composition,13 dysregulating critical host immune responses, and potentially changing important functions of the gut microbiome.15 These effects on the gut microbiome may be strong and persistent, possibly leading to chronic inflammation and tumour progression of even more distal tumour locations.17 However, the association of antibiotics with CRC is complex. Infection is another major risk factor for cancer globally, contributing to approximately 16% of annual incident cancers.19 Some evidence indicates that anti-inflammatory drugs may reduce the risk of cancer,20 further supporting the role of inflammation in carcinogenesis. Thus, it is also possible that antibiotics could reduce the risk of cancer, by decreasing inflammation.Antibiotics have been posed as a risk factor for CRC. Besides the global burden of antibiotic resistance, the high annual antibiotic consumption 21 pooling together colon adenoma with carcinoma22, or combining the different anatomical locations.21 None of them considered departure from linearity, although a recent study suggested for a non-linear dose–response relationship between antibiotic use and CRC risk.9 There is a clear lack of solid evidence summarising the potential association of antibiotics with CRC, and current evidence is insufficient to evaluate whether the potential CRC risk may differ within the colorectal continuum, or by broad- and narrow-spectrum antibiotics, indication, age or sex.22 With the aim of filling these knowledge gaps, we conducted this comprehensive systematic review and dose–response meta-analysis, evaluating the risk of CRC among individuals who ever-used antibiotics.Previous meta-analyses have suffered from power limitations including only five to six studies in total,23This systematic review and meta-analysis was based on an a priori established study protocol. The results are reported in line with the Preferred Reporting Items for Systematic Reviews and Meta-analyses (PRISMA) guidelines.https://www.cochranelibrary.com/central). No restrictions were set for the search.The search strategy consisted of two parts. Firstly, PubMed, Web of Science and Embase were searched from inception until 17th of February 2020 to identify all relevant publications reporting data on the association of oral antibiotics with CRC risk. The search string was initially optimised with the help of Karolinska Institutet University Library  antibiotics with nonusers of antibiotics (0–1 prescriptions), and the risk of primary CRC, colon or rectal cancer.Cohort study, case–control study or randomised controlled trial (RCT).Standardised risk estimates were presented as relative risks (RRs), odds ratios (ORs) or hazard ratios (HRs) with corresponding 95% confidence intervals (CI), or sufficient data were available to calculate these.All data were exported to EndNote X7 and one (JS) author completed the selection based on publication titles. Two independent authors (J.S. and N.B.) selected the publications by assessing the abstracts and full-texts, and any disagreements were solved by mutual consideration between both authors. One author (J.S.) extracted summary data from the publications and another author (R.F.) crosschecked the extracted risk estimates. To be eligible for this meta-analysis, all following criteria had to be fulfilled:Animal and in-vitro studies.Study design: cross-sectional studies, case-reports, abstracts, and reviews, etc.Only pediatric population.Irrelevant exposure or outcome.Following publications were excluded:In case of publications with overlapping data, the latest publication was included. If the publication year was the same, the most comprehensive study was included.24 had scores ranging between 0 and 9, with a total score of respectively, ≤3, 4–6 and ≥7 considered indicative of low, moderate and high quality. Additionally, a customised tool for assessing quality and susceptibility to bias in observational studies was applied.25 Quality assessment was not used to exclude studies.Two independent authors (J.S. and R.F.) completed the quality assessment of the included studies, by systematically applying two quality assessment tools. Newcastle-Ottawa Scale (NOS) tool for cohort and case–control studiesFor each eligible study, we extracted at minimum the following data: (1) study characteristics , and (2) population characteristic .26 providing pooled effect sizes (ES) with 95% CIs for each outcome. Subgroup analyses were performed by anatomical location, study design, antibiotic class and antibiotic group (if reported in two or more studies). Standardised risk estimates  were considered equivalent, as the outcome is rare. The included antibiotic classes were categorised as follows: penicillins, tetracyclines, sulfonamides, macrolides and lincosamides, quinolones, nitrofurans, cephalosporins, carbapenems, nitroimidazole and metronidazole, imidazole and others (based on reported data). Of these, penicillins, metronidazole and lincosamides were considered narrow-spectrum antibiotics, and the remaining broad-spectrum antibiotics.We used DerSimonian & Laird random-effects meta-analysis to pool together the most adjusted risk estimates,27 Departure from linearity was evaluated by Wald test, with a cut off of p < 0.05 indicative of non-linearity. We fitted models with cubic splines and a multivariable random-effects model was used to pool the study-specific estimates.27 To select the best model fit, Akaike information criterion (AIC) was applied.26 The I2 statistics was used to assess statistical across-studies heterogeneity.28 Publication bias and small-study effects26 were assessed using Egger’s test29 and funnel plots.30We performed dose–response analyses to assess the relationship between the number of days exposed to any antibiotics/number of prescriptions and CRC risk. For analyses on the number of prescriptions, non-users were defined as having received 0–1 prescriptions (based on the available data). This could potentially dilute the estimates towards null. We used the median of each prescription category (e.g. 15 prescriptions for category of 10–20 prescriptions) to quantify these as a continuous variable. Firstly, the dose–response model was fitted within each study and variance-weighted generalised least squares (GLS) regression models were used to estimate the pooled study-specific trends.31 (ii) excluding studies without a clear lag-time of at least 1-year, (iii) broad- and narrow-spectrum antibiotics, (iv) excluding/restricting to studies adjusting for non-steroidal anti-inflammatory drug (NSAID), (v) restricting to studies excluding patients with inflammatory bowel disease (IBD), and (vi) restricting to studies excluding patients with Crohn’s disease and ulcerative colitis. All statistical analyses were performed with STATA MP 15.1.To evaluate the robustness of our results several sensitivity analyses were undertaken by: (i) excluding a study with selected population,39 Supplementary Table The search identified 6483 publications, and after the selection process ten studies met the eligibility criteria, including 4,147,560 individuals and over 73,550 CRC cases ,37 five in Europe , one in New Zealand38, and one in Taiwan.31 All publications addressed exposure to oral antibiotics only.39 Six studies39 provided data for cumulative use , and all but two studies36 provided antibiotic class-specific data (for at least 1 antibiotic class). Five studies provided the number of prescriptions,39 four reported the number of days exposed,36 and one study utilised cumulative dose expressed as tertiles.31 Six studies investigated the risk of primary CRC,39 and five studies provided data for colon cancer36 and three studies for rectal cancer.36 All but three studies reported a clear lag-time of at least 1-year.32 The median time elapsed between first antibiotic prescription and cancer diagnosis was 8 years, ranging between 1 and 15 years.Supplementary Table 33 drug registries were used for exposure ascertainment, enabling prospective data collection and eliminating possible recall bias. The outcome was identified from registries or clinical databases in all studies. Although only two studies reported complete coverage of the region (i.e. >90% of eligible participants),36 all studies were population-based and adjusted for age. Apart from two publications,35 the studies clearly excluded antibiotic prescriptions within one year before cancer diagnosis. Additionally, six studies clearly excluded patients with predisposing conditions such as IBD39 or Crohn’s disease and ulcerative colitis.33The quality of the included publications was considered high ,39 whilst no clear association were shown for colon 33 and rectal cancer ,31 based on fewer studies. The statistical heterogeneity was high for CRC (I2 = 95.7%) and colon cancer (I2 = 83.5%), but low for rectal cancer (I2 = 2.7%). Pooling of the four case–control studies yielded similar effect sizes  (Supplementary Table 5).39 The funnel plot was visually asymmetrical towards positive associations, suggesting for presence of small-study effect (Egger’s test p < 0.0005) ,39 whilst no apparent association was shown for studies excluding patients with Crohn’s disease and ulcerative colitis 33 ,34 compared those not adjusting for NSAID use .39The sensitivity analysis restricted to studies excluding patients with IBD suggested for a potentially higher CRC risk ,39 sulfonamides ,38 quinolones ,39 cephalosporins 37 and nitroimidazole and metronidazole 37 (Supplementary Table N = 4).38 No apparent association was found for other antibiotic classes. Supplementary Table The various antibiotic classes (reported in eight publications)Fig. N = ,39 sulfoN = 3)39 among ever-users, whilst no apparent association with was shown for narrow-spectrum antibiotics 39 . The risk appeared to plateau after 30 days of antibiotic use .31 Overall, the quality of included the publications was high, based on the Newcastle-Ottawa Scale and a customised quality assessment tool.The major advantage of this study is the inclusion of over 4.1 million individuals and over 73,550 primary CRC cases, enhancing precision compared to previous meta-analyses or single studies.41 Because of the complexity of the exposure, clinical and statistical heterogeneity were acceptable, yet the sensitivity analyses yielded consistent results.The exposure data were mainly restricted to outpatient care drugs, excluding intravenous antibiotics, as their impact on gut microbiota may differ. However, underestimation of exposure due to over-the-counter drug use is unlikely in highly developed countries, where systemic antibiotics are available only on prescription.43 or for as of yet undiagnosed CRC.44 To minimise potential reverse causality, sensitivity analyses were restricted to publications clearly reporting at least 1-year lag-time,35 and the association remained unchanged.As for all meta-analyses on observational studies, the main limitations of this study are the inability to infer causality and the available data and methods of the included studies. However, to investigate this association by means of a RCT would be challenging, given a large number of individuals should be followed-up over longer time period—limiting feasibility. To see the effects based on real life data, population-based studies offer a favourable and beneficial approach, yet requiring solid research methods and valid data-collection. Another general concern in observational studies is potential residual confounding, and we lacked data to thoroughly assess confounding by indication. Thus, we cannot exclude the possibility that some individuals may have received antibiotics for cancer predisposing conditions, such as impaired immune and inflammation responses,9 This potential association deserves more attention and should be confirmed in other large cohort studies.Data were insufficient to compare the risk of left- and right-sided colon cancer, or to evaluate the effect of indication, age or menopausal status. Some evidence suggests for left sided-colon cancer location, particularly in individuals younger than 50 years.Another limitation is that the pooled risk estimates in this study are a mixture of exposures and exposures periods. Despite that, the majority of the studies had rather similar exposure periods (starting from the 1990’s), the patterns of use, formulations and dosages of antibiotics may have changed over time: related to improved global antibiotic stewardship. Additionally, the duration of exposure periods varied between the studies, complicating the interpretation of the pooled risk estimates. Furthermore, considering the high antibiotic consumption, it is possible that non-users have received antibiotics during a lifetime (i.e. before the study period). However, this left censoring and potential misclassification of exposure would likely be at random among antibiotic users and non-users, potentially diluting the risk estimates towards null, rather than explaining the shown increased CRC risk found for some antibiotic classes.46 It could also lead to an earlier detection of CRC, selectively among antibiotic users,47 overestimating the CRC risk. However, only two of the included studies adjusted for screening colonoscopy.35A risk of detection bias is a concern. Antibiotic users could utilise the healthcare services more frequently, and it is possible that antibiotic users could be more likely to participate in potential screening programmes than non-users. This could result in an earlier detection of colon adenomas, thus significantly reducing the risk of developing CRC.22 this present study is the largest and the first to provide evidence for antibiotic-spectrum specific risk differences, and to investigate non-linear dose–response relationship. We also excluded colon adenomas and divided the colorectal continuum into colon and rectal cancer. However, the association between antibiotics and CRC risk is complex, and this study cannot assess the mechanisms underlying the shown risk differences between the various antibiotic classes. Considering the different mechanisms of action, one could expect differences between the various antibiotic classes, and e.g. quinolones are notorious for DNA-damage. Yet, the microbial dysbiosis observed among CRC patients in some studies,48 supports the biological plausibility of broad-spectrum antibiotics having a long-term effect on the gut microbiota, potentially facilitating colonisation with pathogenic bacteria to a greater extent than narrow-spectrum antibiotics.50 Emerging amount of evidence from human and animal models suggests that dysbiosis, and particularly some bacteria strains , might involve a higher CRC risk through mechanisms increasing cancer promotion by altering cell proliferation, differentiation and apoptosis, inflammation, and production of DNA damaging toxins.52 However, if infection and inflammation can independently promote CRC it is possible that a prolonged exposure to antibiotics could reduce CRC risk, by reducing especially chronic inflammation.18 This could be one possible explanation, or a contributing factor, for the weak non-linear risk pattern shown in this study. However, a prolonged exposure, particularly for indications requiring high exposure to antibiotics, could indicate for a chronic underlying disease. Thus, confounding by indication cannot be excluded, and since fewer individuals are likely to be exposed to high cumulative dosages than to lower dosages, the dose–response data is likely more reliable for those with lower dosages. Antibiotics have previously been associated with an increased risk of several cancer types, including breast53 and lung cancer,42 yet with a decreased risk of ovarian and cervical cancer.54 Although the risk of confounding by indication cannot be ruled out, it may be more of a concern for lung cancer, considering the high use of antibiotics for respiratory tract infections.42 Yet, we lacked data on tumour stage and some of the analyses were limited to fewer studies. Thus, our results should be interpreted cautiously.Compared to previous meta-analyses,In conclusion, this systematic review and dose–response meta-analysis suggest that different antibiotic types involve a different CRC risk. In specific, the excess risk was associated with broad-spectrum antibiotics, whilst no apparent association was found for narrow-spectrum antibiotics. The dose–response analysis did not provide strong evidence of any particular dose–response association, and the risk patterns for colon and rectal cancer were similar. Whereas this study cannot determine causality, the findings raise questions highlighting the need for more mechanistic studies on drug–microbiome interactions, and better understanding of the possible role of antibiotics and microbiome in CRC development.Supplementary Materials


Clusters were chosen to reflect outbreak times  and mobility restrictions (China and Denmark/France/Spain with stricter lockdowns than the UK/US). Results: Overall analysis showed no increase in mortality; however, a statistical increase in mortality was seen in the US/UK cluster with ORp = 1.28  and a decrease in China with ORp = 0.65  and France with OR = 0.31 . Severity and hospitalization were not statistically significant in the analysis; however, several associations were seen in specific countries but not in country clusters. Conclusion: The country-cluster meta-analysis provided a reasonable explanation for COVID-19 mortality among ACEI/ARB users. The analysis did not explain differences in severity and suggested the involvement of other factors. Hospitalization findings among ACEI/ARB users may be considered informative as they may have been subjected to clinical decisions and hospital-bed availability.Background: The use of angiotensin-converting enzyme inhibitors (ACEIs) or angiotensin receptor blockers (ARBs) is controversial for treating COVID-19 patients. We aimed to estimate pooled risks of mortality, disease severity, and hospitalization associated with ACEI/ARB use and stratify them by country and country clusters. Methods: We conducted a search in various databases through 4 July 2020 and then applied random-effects models to estimate pooled risks (OR NPIs may include contact tracing, increased testing, social distancing, wearing masks, and extreme measures such as complete lockdowns and banning public gatherings [entation . The litentation .With the recent concern regarding angiotensin-converting enzyme inhibitors (ACEI) and angiotensin II receptor blockers (ARB), there has been speculation that their use could increase the risk of exacerbating COVID-19 infections by upregulating ACE2 expression ,6. HowevInterestingly, many diseases such as hypertension, diabetes, renal, and cardiovascular diseases share a common indication for ACEI/ARB . Among oBeyond the theoretical risk in these patients, we attempted to evaluate whether or not specific countries and country clusters were successful in mitigating the risk of worse COVID-19 outcomes in ACEI/ARB users compared to non-ACEI/ARB users during their first wave of the pandemic. The results of this study would help to explain several of the discrepancies seen in COVID-19 outcomes associated with ACEIs/ARBs in the published literature. Therefore, the primary objective was to estimate the pooled risks of mortality and disease severity associated with ACEI/ARB use during the first wave of the pandemic. Exploratorily, we also estimated the pooled risk of hospitalization with due caution because of factors independent of COVID-19. We complemented an amended  meta-anaTwo authors  screened publications and extracted odds ratios (OR). For studies not reporting ORs but including adequate data, we estimated the crude OR (unadjusted). Each unadjusted OR was calculated with its standard error and 95% confidence interval according to Altman et al. . In the p) across all studies and then stratified the analysis by studies with and without statistical adjustment. The package uses the generic inverse variance method for the meta-analysis and requires the estimates (OR) and their standard errors as inputs to calculate the pooled estimates in accordance with Borenstein et al. [Using R Core Team (2020) software  and the meta package , we appln et al. .I2 were used to quantify heterogeneity of the included studies. An I2 > 50% indicates significant heterogeneity. We conducted Egger’s test and produced funnel plots to assess publication bias using the same package. A p value < 0.05 considered to be statistically significant. We performed sensitivity subgroup analyses by country and country clusters. The clusters were chosen to reflect time of outbreaks  and mobility restrictions (China and Denmark/France/Spain with stricter lockdowns than the UK/US). Quality assessment was carried out using the Newcastle–Ottawa Scale (NOS) [Q-test and le (NOS) .k) for mortality, disease-severity, and/or hospitalization analyses were included  was statistically non-significant at 0.86 (95% CI = 0.68–1.08) and remaining non-significant when stratified by studies that reported adjusted or non-adjusted estimates (p for COVID-19 disease severity (k = 30) was 0.92 (95% CI = 0.74–1.15) and remained statistically non-significant when stratified by studies with or without adjustments  was 1.17 (95% CI = 0.78–1.75) and remained statistically non-significant in studies with or without adjustment (I2 < 30% = 11) to moderate (2I 30%–60% = 8) with some high (I2 > 60% = 4). Funnel plots and Egger’s tests were significant for mortality (p = 0.04) but not for disease severity (p = 0.216) and hospitalization (p = 0.337), as shown in The single French study reported a statistically significant increase in COVID-19 severity with ACEI/ARB use, while the one British study related a significant decrease in severe COVID-19 disease risk in association with ACEI/ARB use . The ORpjustment . The assjustment . HeterogWith a verifiable total of 43,829 patients in the analysis, including 11,166 exposed to ACEI/ARB, our meta-analysis consistently revealed the absence of an association between ACEI/ARB use and mortality, disease severity, and hospitalization risk in COVID-19, a finding to be validated as further evidence accumulates. The country cluster sensitivity analysis explained several of the differences seen in the mortality outcome. The UK/US cluster revealed an increased risk of mortality. The included studies for this cluster reported non-adjusted OR for the mortality outcome see . The stu2I > 50% in subgroup analysis by countries such as China and US/UK cluster. The included studies defined COVID-19 severity differently and that may explain the seen heterogeneity. Many of the included studies defined COVID-19 severity according to the National Health Commission of China which is similar to the WHO definition [The subgroup sensitivity analyses by country clusters did not explain the heterogeneity in the disease-severity endpoint very well which was evident by finition ,13. The finition . The coufinition . WhetherAn advantage of this current analysis is that we captured published studies from early stages of the pandemic in each country. Therefore, we think that we were able to estimate the risk of mortality during the implementation of NPI measures  in a number of countries. The South Korean national response was a highly successful model for handling the pandemic. With aggressive measures that included contact tracing to prevent community transmissibility, South Korea reported the largest numbers of cases in the first two months of the pandemic. In this current analysis there was no signal of increased mortality among ACEI/ARB users in South Korea . It shouHistory of previous pandemics such The Spanish flu in 1918 taught us that subsequent waves (second and third) is likely to occur. The second wave of the Spanish flu was believed to be caused by a mutated virus . There aIn summary, our meta-analysis of studies accrued to 4 July 2020 suggests no evidence of an association of ACEI/ARB exposure with mortality, COVID-19 disease severity, or hospitalization. The country-cluster meta-analysis provided a reasonable explanation for differences in the mortality outcome, while it failed to explain the severity outcome. More studies of ACEIs/ARBs and COVID-19 severity outcomes in different populations are needed. The association of ACEIs/ARBs with mortality outcomes may be related in part to waves of infection; non-pharmaceutical measures such as mobility restrictions, contact tracing, and increased testing, and progress in vaccines and therapeutic treatments.


PAI-14G/5G (rs1799889) gene polymorphism and the risk of AD, MetS, and female infertility have reported inconsistent results. The aim of the present study was to investigate possible associations.Plasminogen activator inhibitor-1 (PAI-1) is considered to be involved in the physiopathological mechanisms of Alzheimer's disease (AD), metabolic syndrome (MetS), and female infertility. Previous studies investigating the association between Eligible studies were retrieved through PubMed, Medline, EMBASE, CNKI, and WANFANG databases. The odds ratios (ORs) and 95% confidence intervals (CIs) were used to assess the associations. Subgroup analyses by ethnicity and mean age, sensitivity analyses, and publication bias were performed.PAI-14G/5G polymorphism and the risk of AD and female infertility in five genetic models. For the risk of MetS, the PAI-1 4G/5G (rs1799889) polymorphism may be associated with the risk of MetS , especially in Asians  and patients with mean age > 50 years old .Five studies (four articles) for AD, six studies (six articles) for MetS, and four studies (four articles) for female infertility were included in this meta-analysis. Our results showed no significant associations between the PAI-1 4G/5G polymorphism might be associated with the risk of MetS, but no evidence was detected for AD and female infertility.The present meta-analysis suggested that the  Infertility is defined as the failure to conceive after 1 year of regular sexual intercourse, affecting ∼10% to 20% of couples.,7 Of all infertility cases, ∼50% are associated with female factors. Besides environmental risk factors, genetic variation may play a key role in the pathogenesis of AD, MetS, and female infertility.Alzheimer's disease (AD) is a complicated neurodegenerative disorder characterized by vascular amyloid-β (Aβ) deposition, neurofibrillary tangles, and nerve cell death, particularly in some areas of the cerebrum memory and more advanced administrative performance. It plays a critical role in the regulation of the fibrinolytic system by rapidly inhibiting the tissue-type plasminogen activator (tPA) and urokinase-type plasminogen activator (uPA), which cannot cleave plasminogen to give plasmin. The gene coding for PAI-1 has several polymorphic loci, among which the most studied is the 4G/5G insertion/deletion polymorphism containing either 4 or 5(4G/5G) guanine bases at −675 of the PAI-1 promoter.,12 When PAI-1 gene mutations occur, PAI-1 levels will be increased, leading to reduced plasma fibrinolytic activity, which has been observed in a variety of diseases, including AD, MetS, infertility, etc.Plasminogen activator inhibitor type-1 (PAI-1), also known as protein serpin peptidase inhibitor, clade E, member 1 (SERPINE1), belongs to the serine protease inhibitor superfamily.PAI-1 4G/5G polymorphism and the risk of AD, MetS, and female infertility have been investigated by many studies, whereas the results were still inconsistent due to various genetic backgrounds, small sample sizes of each study, and possible biases. To increase statistical power and identify the association between the PAI-1 4G/5G polymorphism and the risk of AD, MetS, and female infertility, a meta-analysis was performed.Up to date, the associations between the 2This meta-analysis was conducted and reported according to the Preferred Reporting Items for Systematic Reviews and Meta-Analyses (PRISMA) 2009 checklist.2.1We searched PubMed, Medline, Embase, CNKI, and WANFANG databases and references of relevant publications only in English and Chinese up to October 2019. Using both medical subject heading terms  and text words . Our study was approved by the Medical Ethics Committee of Shenyang First People's Hospital.2.2Eligible articles had to meet the following criteria:1.PAI-1 gene polymorphisms and risk of AD, MetS, and infertility;articles about 2.sufficient genotype frequencies for estimating odds ratios (ORs) with 95% confidence intervals (CIs); and3.case–control study.Exclusion criteria included:1.insufficient genotype data;2.duplicate studies;3.case reports, reviews, editorials, comments, and abstracts.2.3P-value of Hardy-Weinberg equilibrium (HWE) test in controls, and Newcastle-Ottawa Scale (NOS) score. Two authors independently extracted data and evaluated the study quality using NOS. Total NOS scores ranged from zero to nine stars. A NOS score of 6 to 9 is considered to be of high methodological quality. Discrepancies were resolved through discussion.The following data were extracted from each study: first author, year of publication, mean age, country, ethnicity, genotyping method, number of cases and controls, genotype frequencies in cases and controls, the 2.4PAI-1 4G/5G genetic variants and AD, MetS, and female infertility were reported as ORs with 95% CIs. The ORs and 95%CIs of risks were estimated in allelic, dominant, recessive, heterozygote, and homozygote models for each study. Statistical heterogeneity was quantified using Chi-square-based Q statistics and I2 values. If P > .10 and I2 < 50% indicated significant heterogeneity, fixed-effects models (Mantel-Haenszel method) were used; otherwise, the random-effects models (Mantel-Haenszel method) were merged. The significance of the overall OR was estimated by the Z test. Stratified analyses based on ethnicity and mean age were conducted. HWE was tested in control groups using the Chi-square test. Egger's linear regression test was performed to test for publication bias. If there was publication bias, we recalculated the adjusted ORs using the trim-and-fill method. to evaluate the possible impact of publication bias. Sensitivity analyses were performed by sequentially omitting each study to evaluate the stability and reliability of the results. All statistical analyses were conducted using Stata software . Value of P < .05 was considered as significant difference.In the present meta-analysis, the risks of the associations between 33.1–32 The study screening process is shown in Figure P-values for the HWE and Newcastle-Ottawa scale scores from all included studies are listed in Table Our research yielded 427 studies of potentially relevant studies. After screening, 15 studies (14 articles) were finally enrolled in this meta-analysis.3.2PAI-1 4G/5G polymorphism and the risk of AD, 5 studies (4 articles) including 901 patients and 925 controls were enrolled in this meta-analysis. As shown in Table PAI-1 4G/5G polymorphism was not associated with the risk of AD in five genetic models. Furthermore, stratification by ethnicity failed to explore any association between the polymorphism and AD in Asians and Caucasians.To determine the potential association between the 3.3PAI-1 4G/5G polymorphism and the risk of MetS, 6 studies including 939 patients and 674 controls were enrolled in this meta-analysis. We found a significant association between the PAI-1 4G/5G polymorphism and MetS risk in the overall population under allele , homozygous , and dominant model . Furthermore, stratification by ethnicity identified a significant association between this polymorphism and MetS in Asians under homozygous  and dominant model  (Table P = .029).To determine the potential association between the 3.4PAI-1 4G/5G polymorphism and the risk of female infertility, 4 studies including 1364 patients and 473 controls were enrolled in this meta-analysis. As shown in Table PAI-1 4G/5G polymorphism and female infertility risk in the overall population under any genetic model. Furthermore, stratification by ethnicity failed to explore any association between the polymorphism and infertility in Caucasians.To determine the potential association between the 3.5PAI-1 4G/5G polymorphism and AD risk  compared with unadjusted pooled OR , but this was not significant, indicating that the results were not robust. For female infertility, the trim and fill method revealed no difference between the adjusted pooled OR and 95% CI  and the unadjusted results , which indicated that the presence of publication bias did not influence the stability of the results. Sensitivity analyses showed that several studies had significant impact on corresponding pooled effects, which did not confirm the stability of our meta-analysis (data not shown).Egger's test method was used to assess publication bias. No evident publication bias was detected among the studies on the association of the sk Table . However4PAI-1 4G/5G polymorphism and the risk of AD, MetS, and female infertility. The present meta-analysis suggested that the PAI-1 4G/5G polymorphism might be associated with the risk of MetS, but no evidence was detected for AD and female infertility. Besides the overall analysis, subgroup analyses were also performed according to ethnicity and age. A significantly increased risk of MetS was found in Asians and patients with a mean age > 50 years.We performed a meta-analysis to investigate the association between the  Plasmin, one of the proteases, is known for its role in regulating Aβ levels. Plasmin formation is mediated by plasminogen activators (tPA and uPA) that cleave the plasminogen to the active plasmin. PAI-1 expression inhibits both tPA and uPA and their activity, thus decreasing the level of plasmin. The 4G/5G polymorphism, which is located at the transcription site of the PAI-1 gene, is responsible for increased PAI-1 expression. Thus, the PAI-1 4G/5G polymorphism may be associated with the pathogenesis of AD. It is well known that increased plasma PAI-1 expression and decreased tPA activity are thought to play important roles in the development of MetS. Increased PAI-1 levels have been shown to be associated with many risk factors such as central obesity, hypertension, dyslipidemia, and glucose intolerance, collectively referred to as MetS.,37–39Amyloid plaque deposits composed primarily of Aβ peptides play an important role in the pathology of AD.PAI-1 4G/5G polymorphism in the overall population and subpopulations of AD and female infertility, and could not identify the sources of heterogeneity by stratified analysis based on ethnicity. Third, selection bias may occur due to the inclusion of only English or Chinese literature. Fourth, AD, MetS, and female infertility are multifactorial disorders resulting from complex interactions between genetic, epigenetic, and environmental factors, suggesting that the PAI-1 polymorphism may only partially play a role in the pathogenesis of these diseases, which may result in bias in our results. Finally, our study only determined the associations between a single locus in the PAI-1 gene 4G/5G polymorphisms and three diseases in patients, but we did not examine associations between PAI-1 gene haplotypes and the risk of these diseases due to insufficient haplotype data. It remains unclear whether other PAI-1 gene mutations can lead to changes in its expression. In terms of the genetic causes of disease, haplotypes can provide more critical information than the corresponding single SNP.Several limitations of the present meta-analysis should be considered. First, our study included five studies on AD, six studies for MetS, and four studies for female infertility. The meta-analysis may be unable to have sufficient power to identify real associations because of the limited study number, especially when grouped by ethnicity and age. Second, we observed the heterogeneity of the 5PAI-1 4G/5G (rs1799889) gene polymorphisms might be associated with the risk of MetS, but no evidence was detected in AD and female infertility. However, our results of this meta-analysis should be interpreted with caution owing to the small sample size and small number of studies available. Well-designed studies with large sample sizes in different ethnicities are warranted.In conclusion, the present meta-analysis suggested that We thank all the participants for their cooperation.Conceptualization: Bing Xu.Data curation: Xin Zhang, Bing Xu.Formal analysis: Xin Zhang, Bai Gao.Investigation: Xin Zhang, Bai Gao.Methodology: Xin Zhang, Bai Gao.Project administration: Bing Xu.Resources: Bai Gao.Software: Xin Zhang.Supervision: Bing Xu.Validation: Bai Gao.Visualization: Bai Gao.Writing – original draft: Xin Zhang, Bing Xu.Writing – review & editing: Bing Xu.


CDKN2B/CDKN2B-AS1 have been reported to modulate glaucoma risk in several GWAS across different populations. CDKN2B/CDKN2A encodes tumor suppressor proteins p16INK4A/p15INK4B which influences cell proliferation/senescence in RGCs, the degeneration of which is a risk factor for glaucoma. CDKN2B-AS1 codes a long non-coding RNA in antisense direction and is involved in influencing nearby CDKN2A/CDKN2B via regulatory mechanisms.Variants in Current study investigated four SNPs  of aforementioned genes in a case–control study in a North Indian cohort. Genotyping was done with Taqman chemistry. In addition, an updated meta-analysis was performed.p = 0.045; OR = 0.80(CI = 0.65–0.99) and p = 0.024; OR = 0.73(CI = 0.55–0.96)], respectively than in controls. Genetic model analysis revealed that TT + CT genotype confers 0.73-fold protection against POAG  and trend assumed additive model gives 0.53 times higher protection against PACG progression. However the association of rs3217992 with POAG and PACG did not remain significant after Bonferroni correction. For rs2157719, the ‘C’ allele was found to be less prevalent among cases (POAG/PACG) with respect to controls. Cochran Armitage trend test assuming additive model revealed 0.77 and 0.64-fold protection against POAG and PACG respectively. Bonferroni correction (pcorr = 0.003) was applied and the association of rs2157719 remained significant in PACG cases but not among POAG cases (p = 0.024). The ‘CC’ genotype also confers protection against primary glaucoma (POAG/PACG) among males and female subjects. The frequency rs1063192 and rs4977756 did not vary significantly among subjects, however the haplotype ‘CATA’ was found to be associated with increased glaucoma risk. An updated meta-analysis conducted on pooled studies on POAG cases and controls revealed significant association between rs1063192, rs2157719, rs4977756 and POAG except rs3217992.Two SNPs, rs3217992 and rs2157719 were found to be significantly associated with the disease. The frequency of ‘T’ allele of rs3217992 was significantly lower in cases (POAG/PACG) [INK4 variants and primary glaucoma in the targeted North Indian Punjabi cohort. We believe that deep-sequencing of INK4 locus may help in identifying novel variants modifying susceptibility to glaucoma. Functional studies can further delineate the role of CDKN2B and CDKN2B-AS1 in primary glaucoma for therapeutic intervention.The study concludes significant association between  INK4 locus at chromosome 9p21.3. Cyclin-dependent kinase inhibitor 2B (CDKN2B) and Cyclin-dependent kinase inhibitor 2A (CDKN2A) are two genes located adjacent to each other at INK4 locus in a stretch of about 80 kb. They encode tumor suppressor proteins p16INK4A and p15INK4B, respectively that inhibit cell cycle progression by forming complexes with cyclin-dependent kinase CDK4 or CDK6  and  respectively than control subjects. The frequency of risk allele i.e. C allele of rs3217992 C > T polymorphism was found to be significantly higher in POAG (60.23%) and PACG cases (62.59%) as compared to controls (55.07%) as mentioned in Table p = 0.047; OR = 0.73(CI = 0.54–0.99)] and 0.69-fold  protection against POAG and PACG development and in additive model, TT genotype gave 0.53-fold protection against PACG progression (p = 0.03). However, the association did not remain significant after Bonferroni correction. The distribution of allele frequencies between cases (POAG/PACG) for rs1063192 was not found to be statistically significant as given in Table For genetic association analysis, data was stratified into two groups, POAG cases and PACG cases and comparison was done with controls. Genotype and allele frequencies of the SNPs among cases and controls are given in Table p = 0.058, OR = 0.36; CI = 0.12–1.03). When the comparison in context to mutant genotype (TT) was done among POAG/PACG females with control female subjects, the study failed to obtain any significant difference in the genotype frequency. Genetic model analysis did not reveal a statistically significant difference in genotypic distribution among primary glaucoma (POAG/PACG) males and females as compared to control subjects  but the results were non-significant after Bonferroni correction.Table Genetic data analysis of rs4977756 revealed G allele as the risk allele among POAG cases while A allele was found to confer risk towards PACG, though the distribution of allele frequencies between cases (POAG/PACG) was not found to be statistically significant as given in Table p = 0.0041) and PACG (p = 0.035) males with respect to control males. In the genetic model analysis, recessive model provided 0.29- and 0.22-fold protection against glaucoma (POAG/PACG) progression among male subjects. When the comparison of POAG females was done with healthy female subjects, the study failed to obtain any significant difference in the genotypic frequencies. Genetic model investigation also did not reveal any statistically significant difference in the genotypic distribution among POAG females and control female subjects. In contrast to POAG females, the frequency of CC genotype was significantly different among females having PACG (p = 0.029) with respect to control females. The recessive model unveiled 0.34-fold protection against PACG development in females  but the results were not significant after Bonferroni correction as shown in Table Subsequent segregation of POAG and PACG cases based on gender Table  revealed p = 0.03 males wiCDKN2B  and CDKN2B-AS1  were in very high LD among cases  and controls  respectively. Moreover, rs1063192 was also found to be in strong LD with rs2157719  and rs4977756 . However, controls were observed to be in slightly lower LD as compared to cases for both variants of the two genes for rs3217992/rs1063192  and rs2157719/rs4977756   as given in Table The distribution of haplotype frequency and measure of linkage disequilibrium (LD) for primary glaucoma cases and controls is shown in Table CDKN2B/AS1 genes and POAG. Table Total 125 articles were retrieved by literature survey. After examining all publications, 18 related articles were included along with the current study. Therefore, the current meta-analysis was performed with 19 total studies. This meta-analysis comprised all articles representing total cases  and controls  from 2011 to 2019. All included studies dealt with association between 4 sequence variants of Meta-analysis of rs3217992 comprised of 4 studies including total 2088 affected and 10,266 unaffected subjects. It included 2 studies from Indian population (including the current study). 1 report represented Caucasians and 1 from Japan. All the essential information of included studies is given in Table p < 0.0001, with substantial heterogeneity, I2 = 64.6%. In recessive model (GG vs GA + AA), highly statistically significant association was observed with overall protective effect of GG genotype from disease progression, OR = 0.64; p < 0.00001, with no heterogeneity, I2 = 0% as shown in Fig. 2 = 43.6%) with OR = 0.84 (95% CI = 0.80–0.88) was observed (Table p < 0.0001 to 0.6625 (95% CI 0.5955–0.7369); p < 0.0001. This suggested that exclusion of any single study did not altered the overall combined OR.Main features of results of meta-analysis for rs1063192 are shown in Table p = 0.383) and Egger’s test (p = 0.348) for funnel plot did not reveal any asymmetry in funnel plot ; p < 0.0001 to 0.6625 (95% CI 0.5955–0.7369); p = 0.0002 with 19.2% heterogeneity which differed from pooled OR: 1.2393 (95% CI 0.9612–1.5977), p = 0.0979 with 67.8% heterogeneity  were pooled together to calculate the association of rs2157719 with POAG. It included 1 Caucasian and 4 Asian studies. Out of 4 Asian studies, 2 studies were from Indian population including the current study, 1 report was from China and 1 from Japan as given in Table Total 12 studies were short listed for the meta-analysis of rs4977756 to find out overall genotypic association between this variant and POAG. Total 8739 POAG patients and 21,174 unaffected healthy age and gender matched controls were involved. Meta-analysis contained 3 studies on Caucasians, 4 reports on African ancestry, 5 studies from Asians  as represented in Table p = 0.01, 0.007, < 0.0001 respectively) with pooled OR ranging from 0.46 to 0.77), showing overall protection against POAG as given in Table p < 0.0001) and hence conferring maximum heterogeneity to the data. After omitting this single study there was no heterogeneity (I2 = 0%), suggesting that exclusion of this single study altered the overall combined OR.Main features of results of meta-analysis for rs2157719 are given in Table Visual inspection of the funnel plot did not identify obvious asymmetry. The Egger test for funnel plot asymmetry was not performed as the power of this test was too low to distinguish chance from real asymmetry when the meta-analysis included less than 10 studies.p value (Table p = 0.0005 with very low heterogeneity (I2 = 27.9%) among all included studies in meta-analysis as shown in Fig. 2 = 27.9%. This suggested that exclusion of single study obviously did not alter the overall combined significant association.For rs4977756, the pooled OR was 0.85, under various genetic model  analysis as given in Table p = 0.217) and Egger’s test (p = 0.095) for funnel plot did not reveal any asymmetry in funnel plot as given in Fig. Visual inspection of the funnel plot did not identify obvious asymmetry. The Begg  and CDKN2B-AS1  genes located at the INK4 locus as genetic risk factors for POAG and PACG. CDKN2B is cyclin dependent kinase inhibitor gene that influences cell proliferation and senescence in all cell types   . The ass= 0.75)] , 39, 40. 0.0008) . Yet in  0.0020) . Accordifor POAG . Anothers CDKN2B , 19, 35.CDKN2B-AS1 has been reported to be associated with exfoliation glaucoma in Caucasian population   . The pre8–0.86)] . rs497778–0.86)] . Cao et pulation .p = 0.04) but not in West African population [p = 1.73 × 10−4 [Nominal association of rs4977756 with NTG was observed in African Americans  with POAG and rs10120688 (p = 0.048) in NTG patients (IOP < 21 mm of Hg) thereby suggesting lack of significant genetic association of 9p21 locus with POAG [Previous findings in India and Pakistan have been consistent in terms that no association was observed between glaucoma and K4 locus , 46. A sith POAG . This isCDKN2B rs1063192 and it was found to be associated with decreased risk of glaucoma. Our findings were also in the concordance with previously done meta-analysis [p ≤ 0.0001. Meta-analysis of rs2147719 and rs4977756 also showed statistically significant association of these SNPs with decreased risk of POAG progression with pooled OR ranging from 0.46 to 0.77 and 0.85 to 0.90 respectively, under various genetic models. rs3217992 did not give any association in meta-analysis although found to be associated with primary glaucoma in genetic association analysis in the current study. Whether the association obtained with rs3217992 in the present study represents real cause-effect model relationship as its not supported by meta-analysis should be further investigated. Significant association in some studies could indicate these sequence variants to be a risk factor in certain ethnic groups, which requires more data from different populations to do a meta-analysis and correlate it with ethnicity.Since there is a significant difference in the results of various studied populations worldwide, we did a meta-analysis to measure the overall association between these variants and risk of primary open angle progression. Pooled analysis on POAG cases and controls revealed a significant association between rs1063192, rs2157719, rs4977756 and POAG except rs3217992. Another meta-analysis, conducted by Hu et al.  in 2017 analysis , rs10631INK4 variants significant association was observed with POAG with rs1063192, rs2157719 and rs4977756. Large scale sequencing of the INK4 locus may reasonably detect true functional and rare variants while in vitro and in vivo studies may further assess the functional relevance of these variants in pathogenesis of primary glaucoma.In conclusion, the present study points towards significant association between rs3217992 and rs2157719 with primary glaucoma. The association with rs2157719 with PACG was strong enough to survive Bonferroni correction for multiple testing. In an updated meta-analysis for Additional file 1: Table S1. The Newcastle-Ottawa Scale for the assessment of case-control studies included in the meta-analysis.


Alterations in the levels of arginine and its related catabolic products  in the urea and nitric oxide cycles were reported to play roles in the pathogenesis of major depressive disorder (MDD). The aim of this meta-analysis study is to explore the associations between arginine with its related catabolic products and MDD, and to discuss the possible role of arginine catabolism in the pathoetiology of MDD.This study will be conducted in accordance with the Preferred Reporting Items for Systematic Reviews and Meta-Analyses guidelines. The English language literature published in the databases of PubMed, EMBASE, PsycINFO and Web of Science will be systematically searched. Forest plots will be used to estimate the associations between arginine and its related catabolic products with MDD. Subgroup analysis and meta-regression will also be performed to investigate the source of the potential heterogeneity. Sensitivity analysis will be performed to strengthen the results and to investigate whether any single study would have a significant effect on the results of meta-analysis. Publication bias will be tested for using the funnel plot with Begg test and Egger test. The Newcastle-Ottawa Scale will be applied to assess the risk of bias of observational studies.An integrated assessment of arginine with its related catabolic products may contribute to predict the risk of MDD.The results of associations between arginine with its related catabolic products and MDD will be reported in a peer-reviewed publication. With our findings from this meta-analysis, we hope to provide the most up-to-date evidence for the contributions of arginine and related catabolic products to predict the risk of MDD.https://doi.org/10.17605/osf.io/7fn59].The protocol of current meta-analysis has been registered at the Open Science Framework [Available at:  Accumulating studies have focused on predicting or explaining the onset of MDD and the response of antidepressant treatment through possible biomarkers and their associated metabolic pathway.,5 Arginine catabolism regulation has been received increasing attention due to dysfunction of oxidative and nitrosative stress in individuals with MDD.–8Major Depressive Disorder (MDD) is 1 of the most common psychiatric disorders and leading causes of global burden of disease, which affected more than 350 million people worldwide.,10 , and glutamate in prefrontal cortex of brain, which are primarily considered play important roles in the cellular bioenergetics and oxidative stress. Additionally, 2 clinical trials indicated that ketamine and esketamine contributed to the alterations of arginine and other metabolites in the urea cycle.,22 Since accumulating evidence has revealed the associations between arginine and its related catabolic products in the urea and NO cycles with MDD, these findings support the role of these metabolites as potential putative diagnostic markers of MDD. However, studies evaluating arginine and its related catabolic products have not clearly illustrated the change pattern between the levels of these metabolites and the MDD.Arginine, a semi-essential amino acid, is reported as a substrate for important pathways to physiological processes in the central nervous system and immune defense such as urea and nitric oxide (NO) cyclesThe aim of this study is to explore the alterations of arginine and its related catabolic products  in individuals with MDD. The comparisons will be conducted between individuals with MDD and healthy controls . We hypothesized that an integrated assessment of arginine with its related catabolic products may contribute to predict the risk of MDD.22.1 Preferred Reporting Items for Systematic Reviews and Meta-Analyses flow diagram of study selection process is shown in Figure https://doi.org/10.17605/osf.io/7fn59]. The search strategy of PubMed is shown in Table This systematic review and meta-analysis will be conducted in accordance with the Preferred Reporting Items for Systematic Reviews and Meta-Analyses.2.2The case-control or cohort studies meeting the following inclusion criteria will be included in the analysis:(1)adult subjects (≥18 years old);(2)assessed a group of unaffected ‘controls’, control subjects were selected from healthy volunteers who were recruited from students, and company employees and so on documented to be free from psychiatric problems and histories of mental illness;(3)eligible studies included participants meeting the Diagnostic and Statistical Manual of Mental Disorders (no restrictions on editions) or International Classification of Diseases and Related Health Problems criteria for depressive disorders;(4)the study measured arginine and/or ornithine and/or citrulline and/or argininosuccinate levels among all participants; and(5)the levels of arginine and/or ornithine and/or citrulline and/or argininosuccinate of participants in the MDD case group and healthy control group were available in the study.Exclusion criteria of the studies were met if they(1)did not focus on evaluating the arginine or ornithine or citrulline or argininosuccinate levels on MDD;(2)only included unhealthy control group with other diseases ; or(3)were repetitive publications from the same datasets by the same or different authors.2.3The outcome for this meta-analysis will be the differences of arginine and its related catabolic products  between individuals with MDD and healthy controls, as measured by the standard mean differences (SMDs) of their concentrations. Paired investigators will independently select the studies, reviewed the main reports and supplementary materials, extracted the relevant information. All reference lists of the retrieved articles will be reviewed to identify the potential studies. The following information will be extracted from each study: first author, publication year, study design, country, geographic location, age, sex, body mass index, type of blood sample specimen required for test, sample detection method, sample size, subjects’ mean arginine or ornithine or citrulline or argininosuccinate levels, and standard deviations.2.4I2) test, which shows that the percentage of the variability in effect estimates owes to heterogeneity rather than chance. If P < .10 or I2>50%, we consider that the heterogeneity had statistical differences and we will use a random effects model. Otherwise, the fixed effect meta-analysis will be applied. According to the statistical power analysis for the behavioral sciences (2nd edition), the effect size (ES) of SMD is judged using the following rules: ’trivial’ (ES < 0.20), 'small’ (0.20≤ ES <0.50), ’medium’ (0.50≤ ES <0.80), and a ’large’ effect (ES≥0.80)”. . Subgroup analysis will be performed to explore the potential influence of included characteristics of the studies on the pooled ES. Meta-regression will also be performed to investigate the source of the heterogeneity, and the effect of both continuous and categorical factors on the study can be assessed simultaneously. The subgroups will be created according to mean or median age, publication date, mean or median body mass index, geographic location, type of blood sample, sample detection method, and so on.All the data analyses will be conducted using Stata . Forest plots will be used to estimate the association between arginine and its related catabolic products with MDD, which will be evaluated by SMD with a 95% confidence interval. The heterogeneity across the studies will be evaluated by chi-square statistics and I-Squared ( The NOS include 3 aspects: participant selection (0 - 4 points), comparability of the study groups (0 - 2 points) and the assessment of the outcome or exposure (0 - 3 points). The total NOS scores categorized into 3 groups: low risk of bias (7 - 9 NOS points), high risk of bias (4 - 6), and very high risk of bias (0 - 3). All 2-tailed P-values < .05 will be defined as statistical significance.Sensitivity analysis will be performed to strengthen the results and investigate whether any single study would have an effect on the heterogeneity of total measurements in each meta-analysis. The funnel plot with Begg test and Egger test will be used for testing the publication bias. The Cochrane Collaboration recommended the Newcastle-Ottawa Scale (NOS) as an assessment tool for testing the risk of bias for observational studies.2.5No patients will be involved in this study.2.6All analyses were based on previous published studies, thus no ethical approval and patient consent are required.33.1There are growing interests in the researches of arginine catabolism/NO regulation in the development and progression of MDD. Our current meta-analysis is expected to clarify the associations between 4 metabolites , and furtherly illustrate the correlations among these metabolites in the NO regulation pathway of MDD in Figure 3.2Due to the following potential limitations, the results of our meta-analysis may need to be interpreted carefully. First, we might not find a sufficient amount of original researches to perform the analyses. Second, we are focusing on 4 metabolites  and perhaps there are not enough reports of each metabolite. Third, our meta-analysis will only include the full-length articles published in English language journals. The data from gray literature (such as unpublished studies and non-peer reviewed literature) or non-English language journals will be excluded. Thus the number of studies might be quite small. Additionally, the potential high heterogeneity caused by differences in age, sex, geographic location, racial, and so on, will lead to selection bias and also decrease the reliability of our results.To the best of our knowledge, this will be the first meta-analysis exploring the links between the arginine with its related catabolic products and MDD. With our findings from this meta-analysis, we hope to provide the most up-to-date evidence for the contributions of arginine and related catabolic products to predict the risk of MDD.None.Conceptualization: Bing Cao, Xiao Gao.Data curation: Bing Cao, Runze Deng, Li Li.Investigation: Li Li, Runze Deng.Methodology: Bing Cao, Dongfang Wang.Software: Bing Cao, Li Li.Writing – original draft: Bing Cao, Runze Deng, Dongfang Wang, Li Li.Writing – review & editing: Xiao Gao, Zhongyu Ren, Lixin Xu.


While it has been claimed that lung cancer occurs due to epigenetic mechanisms, four systematic reviews were reported to investigate the association between serum folate levels and lung cancer risk. Considering some methodological problems founded in the systematic review, a meta-epidemiological study was conducted. The selection criteria of this study were defined that a case-control study was conducted to determine the risk of lung cancer occurrence according to the concentration of serum folate and its results showed odds ratio and its 95% confidence interval. Additional paper was explored from cited lists of 4 papers selected by previous systematic reviews. Random effect model was applied if I-squared value was over 50%. For 5 case-control studies selected, the summary odds ratios  were 0.82 (0.74-0.90) in men, 0.70 (0.62-0.79) in former smokers, and 0.86 (0.75-1.00) in non-smokers. Higher foliate levels can decrease lung cancer risk in men and former smokers. Especially, the protective effect was highest in former smokers compared in non-smokers and current smokers. Based on these facts, folate fortification programs to reduce lung cancer risk would be focused on former smokers in men. And some epidemiological studies are needed to provide a hypothesis to explain the sex differences in the association between folate and lung cancer risk. Lung cancer ranks first in incidence and mortality rates worldwide , and smoking is the best-known risk factor . Recently, it has been claimed that lung cancer occurs due to epigenetic mechanisms caused by exposure to air pollution .9) is known to be involved in the DNA methylation process , 4 systematic reviews were reported to investigate the association between serum folate levels and lung cancer risk  , and refers to DNA methylation, histone modification, small RNA, and so on in the aspect of cellular molecule . While folate  . Two sysAccordingly, it is necessary to conduct another meta-analysis using odds ratio with applying more stringent selection criteria. In order to re-investigate the association between serum folate levels and the lung cancer risk, a meta-epidemiological study was conducted . As the main aim of meta-epidemiological study is to evaluate problems associated with errors that can occur while performing a systematic review , the subjects of analysis in meta-epidemiological study are original articles selected in systematic reviews . Considering previous systematic reviews’ problems pointed out in introduction, the selection criteria of this meta-epidemiological study were as follows. A case-control study was conducted to determine the risk of lung cancer occurrence according to the concentration of serum folate and its results showed odds ratio and its 95% confidence interval (95% CI). Applying the selection criteria, a total of four case-control studies  were selected among the papers selected from the 4 systematic reviews presented in In each of the finally selected papers, the information to be used for meta-analysis was extracted in two ways. The odds ratio of the highest category and its 95% CI value were taken as information extracted by the highest versus lowest method. And the interval collapsing method was applied to utilize the information of all categories . In other words, a meta-analysis of the results of the other categories except for the reference category was carried out, and then the summary odds ratio was counted as information extracted by the interval collapsing method of the paper. Logarithm odds ratio and its standard error were calculated from the extracted information.The heterogeneity level was evaluated by the I-squared value (%), and the meta-analysis was applied to the random effect model for 50% or more of I-squared value . Subgroup analyses were performed according to the sex – men and women – and smoking habits – non smokers, former smokers, and current smokers. Egger test was conducted to check for publication bias . Statistical significance level was calculated as 0.05.As of December 31, 2019, a total of 90 papers cited 8 papers . When the selection criteria were applied, one case-control study published in 2019 could be added . Therefore, 5 case-control studies were finally selected for meta-analysis . The results of the study showed that the higher the serum folate level in men, former smokers, and non smokers decreased the risk of lung cancer occurrence. The summary odds ratios and their 95%CI of subgroup analyses were as same as Yang et al., (2018) except for current smokers. Although this study applied more stringent selection criteria than Yang et al., (2018), it is assumed that the database applied to the analysis of subgroups by sex and smoking habit was the same. Of the 14 papers selected by Yang et al., (2018), 3 papers  met the selection criteria in this study, and their Newcastle-Ottawa Scale was 8 or more. The lower I-squared value from 89.4% to 68.6% could be deduced by applying more stringent selection criteria. From these findings, it can be reaffirmed that suggestion of valid selection criteria in the planning process and the strict application in the evaluation process are particularly important issues in conducting a systematic review.In a systematic review of nutritional epidemiology studies that present disease risks according to their distribution in a particular food or nutrient, the highest versus lowest method has a limit to ignore some information . In both the results of each paper  as well The level of folate in nutritional epidemiology is determined by measuring serum and RBC concentrations, using the level of intake through questionnaires, or using the dose administered during trials . Since the serum concentration reflects a recent uptake , this study selected case-control studies that measured serum folate. Interestingly, systematic reviews of prospective cohort studies reported the same result that there was no link between intake of folate and lung cancer risk . In particular, the subgroup analysis of dose-response meta-analysis  showed the same result that the protective effect was found in men. In other words, different measures of serum concentrations and intake levels for folate were applied, but the same conclusion was found to be protective in men.Sex differences in the association between folate and lung cancer could be considered as follows. First, to date, there is only one database that examined the incidence of female lung cancer according to serum foliate levels . But, ChIn conclusions, higher foliate levels can decrease lung cancer risk in men and former smokers. Especially, the protective effect was highest in former smokers compared in non-smokers and current smokers. Based on these facts, folate fortification programs to reduce lung cancer risk would be focused on former smokers in men.


The risk of prostate cancer in melanoma patients has been frequently assessed. However, a comprehensive meta‐analysis specifically examining this association is lacking. Our aim was to quantify the risk of prostate cancer in melanoma patients based on the available evidence.A systematic review of the existing literature was performed in PubMed, Scopus, and Cochrane databases by two authors independently. Studies reporting the effect size in the form of standardized incidence ratio (SIR) were used for quantitative analyses.I2 = 75%).Of 17 studies included in the review, a total of 15 studies with 282 592 male melanoma patients were used for the analysis. Random‐effects meta‐analysis found a 24% increased risk  of prostate cancer in melanoma patients compared to the general population, with a prediction interval of 1.05 to 1.45. The risk was consistently significant for various geographical regions and latitude. Heterogeneity was significant (These results suggest that the incidence of prostate cancer is significantly higher in patients with melanoma compared to the general population. Our study findings show that melanoma patients have a 24% higher risk of developing prostate cancer. This could be attributed to various environmental, hormonal and genetic factors. Early diagnosis of prostate cancer in melanoma patients may help in risk stratification. A standardized data‐collection protocol was followed to gather the data from each study and included the primary outcome with the number of observed and expected cases, comparison population, publication year, country of study, study period, study design, adjustment factors and method of diagnosis of melanoma and PC. Both authors assessed the quality of the studies using the Newcastle‐Ottawa scale (NOS) for cohort studies. Any discrepancies between the evaluators regarding the inclusion of the studies were resolved by consensus.2.1I2 statistic were used to assess the heterogeneity. I2 value greater than 50% was considered to denote significant heterogeneity not attributable to random error.P‐value of less than .05 was considered statistically significant for all analyses. We used Comprehensive Meta‐Analysis software  for all statistical analyses and generation of graphical plots.The reported SIRs and corresponding 95% CIs were used when provided. For studies which reported the number of observed cases, expected cases, and SIRs, the 95% CI was calculated assuming a Poisson distribution. For studies with fewer than five observed cases, SIRs were not calculated due to the risk of instability.3The initial search identified 1351 records which were screened according to the inclusion and exclusion criteria. Seventeen studies were finally included in the study of which 15 studies were used for quantitative analysis.3.1Seventeen studies reported either SIR or the number of observed and expected cases. Two of these studies reported fewer than five observed cases and were excluded from the meta‐analysis.3.2I2 = 75%, Figure The random‐effects meta‐analysis of 15 studies revealed 24% increased risk of prostate cancer in melanoma patients compared to the general population, with SIR of 1.24  showed significantly increased risk  of PC in melanoma while the smaller studies showed a statistically nonsignificant increased risk . Similarly, this effect estimate was significant  for registry‐based studies and nonsignificant  for single‐center cohort studies. Analyses according to geographical regions and latitude showed statistically significant results in all subgroups. The results of subgroup analyses are summarized in Table 3.4P = .18) and Egger's test (P = .28) did not show any evidence of publication bias.Visual examination of the funnel plot to assess the publication bias showed symmetry and absence of bias Figure . Begg's 4This study found that the melanoma patients were at higher risk of developing subsequent PC than the general population. The results in this study were mainly pooled from population‐based registries across various geographical regions. A previous meta‐analysis which assessed the risk of various secondary primary cancers in melanoma patients had also reported an increased risk of PC in patients with melanoma .The precise pathophysiological mechanism responsible for the increased risk of PC among patients diagnosed with melanoma is currently unknown. Researchers have attempted to explain the association using various hypotheses pertaining to environmental, hormonal, and genetic aspects, among which, the role of sun exposure is the most described one. Exposure to solar radiation, especially intermittent sun exposure and a history of sunburn is considered a leading environmental cause of melanoma.Androgen dependency of both melanoma and PC may explain the relationship between the two tumors. The role of androgens in the carcinogenesis is well known. Androgens have also been found to play an important role in the development of melanoma through multiple mechanisms which include increased melanoma cell proliferation, stimulation of telomerase activity, and suppression of the immune system.Detection bias is an important factor to consider while analyzing the results of epidemiological studies. Detection of one type of cancer may lead to the increased surveillance for the cancers occurring elsewhere. Similar phenomenon may have played a role for the increased risk of PC in melanoma patients seen across the included studies. Except for the study by Kok et al,In subgroup analyses, we found that larger studies and registry‐based studies  showed significantly increased risk while smaller and single‐center cohort studies showed nonsignificant risk. Subgroup analyses also showed higher risk in European studies compared to the studies from Australia and North America; however, the confidence range was comparatively wider for the analysis, which limited the possibility of deriving any conclusions. Although we did not perform a formal analysis due to limited data, we observed that the risk of developing PC was lower for head and neck melanoma compared to the melanomas at other sites.I2 value as a sole measure of heterogeneity has been questioned and prediction interval is recommended to assess heterogeneity in random‐effects meta‐analyses.The limitation of our study include a considerable heterogeneity due to the variation in effect estimates of the included studies, which could be attributed to the baseline differences among included subjects. We attempted to address this limitation through various approaches. First, we performed subgroup analyses according to various categorical variables. We used a random‐effects model for the analysis and also calculated the prediction interval. The role of This study based on available evidence showed an increased risk of PC in melanoma patients. However, it is still unclear whether melanoma is a causal risk factor for PC or weather the detection bias resulted in the increased incidence of PC in melanoma patients. Future studies should consider assessing whether the melanoma patients actually benefit from screening for PC.The authors declare no conflicts of interest.PA and MM contributed to study design and literature review and analyses.PA contributed to literature search and analyses. MM contributed to manuscript drafting.