Breast cancer (BrCa) is a heterogeneous disease and has important variability according to ethnicity and race with respect to incidence, clinical characteristics, and prognosis. Numerous epidemiological studies indicate that BrCa and it’s also related to environmental factors. We, therefore, undertook a systematic review of the literature regarding BrCa risk in women who used OCs based on case-control studies carried out in the years 2009–March 2020 and then performed a meta-analysis of relevant data. Increased BrCa risk was associated with early menarche, nulliparous, non-breastfeeding, older age at first parity, postmenopause, obesity, smoking, and family history of BrCa.p = 0.0358. Regarding other risk factors for BrCa, we found that increased risk was associated significantly with early menarche, nulliparous, non-breastfeeding, older age at first parity, postmenopause, obesity, smoking, and family history of BrCa. Despite our conclusion that birth control pills increase the cancer risk being supported by extensive previous studies and meta-analyzes, further confirmation is required.To perform a meta-analysis of case-control studies that addressed the association between oral contraceptive pills (OC) use and breast cancer (BrCa), PubMED (MEDLINE), Embase, and the Cochrane Library were searched to identify case-control studies of OC and BrCa published between 2009 and 2020. We used the DerSimonian–Laird method to compute pooled odds ratios (ORs) and confidence intervals (CIs), and the Mantel–Haenszel test to assess the association between OC use and cancer. Forty-two studies were identified that met the inclusion criteria and we included a total of 110,580 women . The conducted meta-analysis showed that the use of OC was associated with a significantly increased risk of BrCa in general, OR = 1.15, 95% CI: 1.01 to 1.31,  Among malignant tumors, BrCa is the leading type of cancer that affects women in most countries. According to the predictions of the WHO’s International Agency for Research on Cancer, the specific age-standardized incidence rate for cancer of the female breast worldwide is 46 cases per 100,000 women. Herein, incidence rates are elevated in Australia/New Zealand (94), Western Europe (93), Northern Europe (90), and North America (85). In contrast, incidence rates in sub-Saharan African regions, particularly in Eastern (30) and Middle Africa (28), as well as South-Central Asia (26), are considerably lower , compared with never-users. There was no significant increase in risk among premenopausal women , postmenopausal women , or nulliparous women  [The presented meta-analysis of the magnitude of effects of OC on BrCa risk revealed a statistically significant slight increase of cancer risk, OR = 1.15 . The results from our study were, to a large extent, consistent with the results described in previous studies. For example, in 2018, 1245 BrCa cases were found in the analysis of data derived from a prospective national population-based cohort study carried out from 1991 to 2007 of 74,862 Norwegian premenopausal women. These authors saw a statistically insignificant increased risk of BrCa in participants using hormonal contraception. Here, HR (hazard risk) = 1.12, 95%: CI 0.99 to 1.26 . Moreove82–1.26) .Gierisch et al.  also conThe results of our meta-analyses of potentially modifying risk factors for BrCa indicate that menarche before the age of 12 years, nulliparous, non-breastfeeding, first pregnancy in old age, postmenopause, obesity, smoking, and family history of BrCa significantly increase the risk of development of BrCa. The majority of the above risk factors are associated with a longer duration of estrogen exposure ,77,78,79We are aware that drawing final conclusions from the results of our meta-analysis requires caution given the number of limitations encountered in its construction. First, we limited the search to studies published in English, and these were identified through electronic databases. The possibility of not reaching out to all publications on this topic may have an effect on the value of the results . SecondlIn conclusion, the results from this meta-analysis suggest that OC using is associated with a modest, statistically significant increase in the risk of BrCa. Despite our conclusion that birth control pills increase cancer risk, and that our conclusion is supported by extensive previous studies and meta-analyzes, further confirmation is required.


Many studies have described the relationship between kidney stones and stroke, but the results are controversial, so we conducted this meta-analysis to estimate the relationship between kidney stones and the risk of developing stroke.I2. We performed subgroup and sensitivity analysis to assess potential sources of heterogeneity.Studies were marked with a comprehensive search of PubMed, EMBASE, Google, and ISI Web of Science databases through 25 March 2020. Hazard ratios (HRs) and 95% confidence intervals (CIs) were extracted, and a random-effects model or fix-effects model was used to compute the pooled combined risk estimate. Heterogeneity was reported as I2=79.6%; p=0.000). We conducted a sensitivity analysis by removing the studies that had a high risk of bias. Heterogeneity subsequently decreased significantly, while an increased risk of stroke in patient with kidney stones was again demonstrated . Stratifying analysis showed that the results were more pronounced for ischemic stroke  and the follow-up duration ≥10 years .Eight studies of seven articles involving 3,526,808 participants were included in the meta-analysis. Overall, kidney stones were associated with a moderate risk of stroke incidence (HR, 1.24; 95% CI, 1.11–1.40; Our meta-analysis suggests that patients with kidney stones may have a modestly increased risk of developing stroke, especially in ischemic stroke. More large-scaled and clinical trials should be done to identify the relative impact of kidney stones on stroke outcomes in the future.The online version contains supplementary material available at 10.1007/s10072-021-05113-5. Stroke is a common cerebrovascular disease known for its high incidence, high mortality, and increased disability rate. Depending on statistics, it is the second-largest cause of death and disability in the world –3. On avKidney stones are a relatively prevalent disease in our daily life, especially in Western civilizations , 7. ManyOur meta-analysis was conducted and reported according to the recommendations of Preferred Reporting Items for Systematic Reviews and Meta-Analyses (PRISMA) Statement . We searStudies were included for our meta-analysis if they fulfilled the following criteria: (1) the study had a cohort design, case control, or cross-sectional design; (2) assignment of kidney stones as the baseline exposure, stroke was the outcome measure; and (3) the reported quantitative estimates of the multivariate-adjusted hazard ratios (HRs), odds ratio (OR) or relative risk (RR), and 95% confidence intervals (CIs) for stroke associated with kidney stones. The original effect size value  was used 1:1 as input for HR in the sense of best estimation.All data were independently abstracted in duplicate using a standard data collection form. When necessary, the original authors were contacted for supplementary information. Discrepancies were settled by consensus. The following data were extracted from each study: first author’s last name, year of publication, study design, source of population, the country where the analysis was performed, size of the study, age range or average age, follow-up time, assessment of kidney stones and stroke, and study quality. The Newcastle–Ottawa Scale (NOS) was utilized to assess the quality of the studies . The fulP values were 2-sided, and p < 0.05 was considered statistically significant. For each study, the HR was used for the measurement data and presented with 95% confidence intervals. The chi-square test’s degree of heterogeneity among the results was estimated . Whenever significant heterogeneity  was achieved, the random-effects model was used. If no significant heterogeneity across studies was found, a fixed-effect model was selected to pool the data [Stata 12  was utilized to perform the meta-analysis. the data . The pStn=5) [n=5) [n=2) [A flow chart displaying the results of the literature search and selection is presented in Fig. n=5) –24, no s5) [n=5) –29, and 5) [n=2) , 31. Sev5) [n=2) –13, 16.FThe characteristics of the studies and their participants are presented in Table I2=79.6%), so we used the random-effects model. Multivariable-adjusted HR of stroke incidence concerning kidney stone from individual studies is presented in Fig. p = 0.000). When we excluded the cross-sectional study [p = 0.001. The result was statistically significant, confirming the increased risk of stroke in patients with kidney stones again.Eight studies of seven articles –13, 16 ep = 0.014; North American: 1.26 , while Europe studies were not associated with a risk of stroke . The pooled estimate of multivariate HRs based on three studies [p = 0.000) among ischemic stroke and 1.07 among hemorrhagic stroke  based on one study [There was statistically significant heterogeneity in the overall analysis, and the source of population, duration of follow-up, study design, and methodological quality varied across studies. Therefore, we also conducted subgroup analyses further to elucidate the effect of kidney stones on stroke risk. Table  studies , 12, 13 p=0.000), the heterogeneity was significantly decreased   and Egger’s test (p=0.929).We tested the robustness of our results in a sensitivity analysis by omitting one study at one time. The results showed that the study of Chung et al.  substant09) Fig. . There w09) Fig. . FurtherOur meta-analysis from eight studies of seven articles confirmed a positive association between kidney stones and the stroke risk after adjusting established cardiovascular risk factors, with an overall 1.24-fold increased risk compared with those without a history of kidney stones.Over the past decades, despite extensive studies investigating the kidney stone’s role on either cardiovascular disease or stroke, it remains unclear whether the association between kidney stones and risk of stroke incidence is causal. Some of the studies suggested that kidney stones were associated with an increased risk of stroke, and the others failed to find the association. Many studies showed that kidney stones are associated with coronary heart disease (CHD) , 32, hypFor the moment, the mechanisms underlying the increase of consequent stroke events in kidney stones patients are not well understood. Previous studies showed that kidney stones are associated with hypertension, smoking, obesity, diabetes, and hyperlipidemia, which are all known risk factors for stroke , 34, 36,I2=28.7%; p=0.000), we found that the study was a prospective cohort study, which was conducted by the source from the Taiwan Longitudinal Health Insurance Database 2000. The diagnosis of kidney stone and stroke in this article depends on the doctors’ and hospitals’ data, rather than based on standardized criteria, and residual confounding factors and surveillance biases also existed. As we all know, according to the GRADE system, five factors may lead to a decrease in the quality of evidence, including the risk of bias, inconsistent results, indirectness of evidence, inaccuracy of the results, and publication bias. In this article, we believe that the inconsistency in rating the quality of evidence is the main reason for the current results and heterogeneity. Of course, differences in environmental factors, countries, methodological factors in study design, and how the studies were conducted should attribute to the observed heterogeneity. Therefore, the existence of heterogeneity requires a proper and careful understanding of the current meta-analysis findings.However, heterogeneity between studies was found in kidney stones and risk of stroke incidence. Therefore, we chose the random-effects model to analyze the results; stratifying analysis and sensitivity analysis were also used to reduce heterogeneity and find the source of heterogeneity. Within the stratifying analysis, we examined study design, geographical area, stroke type, length of follow-up, and study quality as possible sources of heterogeneity, and some studies showed that these did not show any significant heterogeneity between studies. Although the stratifying analysis could not explain the heterogeneity level in interpreting the results, several studies’ differences are worth discussing. When we conducted a sensitivity analysis by omitting one study at one time, the results showed that the study of Chung et al.  substantThis study has several important strengths compared with previous meta-analyses. To our knowledge, this article provides a more systematic explanation of the relationship between kidney stones and stroke risk. It also conducts stratifying analysis on some known influencing factors. At the same time, sensitivity analysis is used to screen the research results’ heterogeneity. Furthermore, the Newcastle-Ottawa Scale was used to assess individual studies’ quality. Most of them were of high quality. Besides, we use the funnel chart to directly show that there is no publication bias in the research results, which are further confirmed by Begg’s and Egger’s tests. Therefore, the results should be reliable.There were also many limitations to our meta-analysis. Firstly, a limited number of studies were included, although we strove to find all related studies. Only eight studies of seven articles were included in our meta-analysis. Secondly, although we considered a multitude of risk factors in the multivariable analysis, the possibility of residual confounding or confounding by unmeasured factors, which cannot be ruled out in any observational study, must be acknowledged. Thirdly, although we used the random-effects model and stratifying analysis, the results still have some heterogeneity. Also, the number of studies included in the subgroup was small, lacking sufficient reliability to confirm a relationship in a definitive manner. The number of available studies of different stroke outcomes and differences in study design, country, and assessment method included in this meta-analysis was moderate. Therefore, the results could be influenced by factors like methodological differences, regional differences, etc.Our meta-analysis found a moderate association between kidney stones and the risk of stroke incidence after adjustment of established cardiovascular risk factors, especially in ischemic stroke. To efficiently assess the association and causality of kidney stones on stroke, more large-scaled and clinical trials should be done to identify the relative impact of kidney stones on stroke outcomes and explore whether effective treatment of kidney stones may prevent or improve the course of stroke.ESM 1(DOC 60 kb)


SERPINE1 whose the role in the risk of developing DN remains questionable. In order to elucidate the contribution of SERPINE1 in DN progression in the context of type 2 diabetes mellitus (T2DM), we conducted an association study and meta-analysis of SERPINE1 genetic variants. Materials and Methods: A total of 190 patients with DN, 150 T2DM (type 2 diabetes mellitus) patients without DN and 238 healthy controls were recruited. We selected five tag single-nucleotide polymorphisms (SNPs) from the HapMap. The generalized odds ratio (ORG) was calculated to estimate the risk on DN development. Subgroup analyses based on ethnicity and type of diabetes were also performed. Results: Both the present association study regarding SERPINE1 SNPs  did not found any significant association between SERPINE1 variants and DN and the meta-analysis of variant 4G>5G (rs1799889) did not also reveal a significant association between 4G>5G variant and DN in main and subgroup analyses. Discussion: In conclusion, the present association study and meta-analysis provides strong evidence that SERPINE1 genetic variant 4G>5G is not implicated in the risk or development of DN in Caucasians. Further studies in other populations remain to further investigate the role of this variant in the course of DN.Background: Many lines of evidence highlight the genetic contribution on the development of diabetic nephropathy (DN). One of the studied genes is  Diabetic nephropathy (DN) constitutes one of the microvascular complications of both type 1 diabetes mellitus (T1DM) and type 2 diabetes mellitus (T2DM) and the most common form of chronic kidney disease . Many geTwo main approaches were used for the genetic dissection of the DN, genetic linkage studies and genetic association studies ,7. ThereExcept individual studies, there are also available meta-analyses of these studies. The most recent meta-analysis of genome-wide linkage studies (GWLS) identified significantly suggestive for linkage with DN cytogenetic locations on the following chromosomes: 1q, 3q, 4p, 5q, 7q, 15q, 16p, 17q, 19q and 22p . On the SERPINE1) gene, also known as Plasminogen activator inhibitor-1 (PAI-1), which is a member of the serine proteinase inhibitor (serpin) superfamily and constitutes the principal inhibitor of tissue plasminogen activator (tPA) and urokinase (uPA) leading to inhibition of fibrinolysis  and data for nominal variables as count (or ratio) and percentage [n (%)]. The association between genotype distribution and disease was examined using the generalized odds ratio (ORG) ,40. In hG was calculated using ORGGASMA (http://biomath.med.uth.gr accessed on 30 August 2021) [ORst 2021) ,40. StatSERPINE1 or “Plasminogen activator inhibitor-1” or PAI-1) AND  AND (genetic or association or gene or polymorphism). We also retrieved articles from genome-wide association studies (GWASs) Catalog (https://www.ebi.ac.uk/gwas/ accessed on 1 December 2020). All of the references cited in the included studies were also scrutinized to identify additional published work. Meta-analyses of the included genes were also screened. Case reports, editorials, and review articles were also excluded. The search was restricted to articles in English.All of the studies published until December 2020 were identified by extended computer based search of PubMed database. The following search terms were used:  [p < 0.10) [2 statistic [G was calculated using a software for implementing the generalized odds ratio methodology for the analysis and meta-analysis of GAS (ORGGASMA) (http://biomath.med.uth.gr accessed on 30 August 2021).The association between genotype distribution and disease progression was examined using the generalized odds ratio (ORG) ,40. The G)  random e < 0.10)  and we atatistic . ORG wasFor each study, we examined whether controls confronted with Hardy–Weinberg equilibrium (HWE) predicted genotypes using Fisher’s exact test. For studies providing only allele counts, we relied on the authors’ assessment of deviations from HWE. We also tested for small-study effects with the Egger test .The cohort consisted of 190 cases (patients with T2DM and DN), 150 diseased controls (patients with T2DM without DN) and 238 healthy controls. All patients were Caucasians of Greek origin. The demographic and clinical characteristics are shown in G, are shown in p ≥ 0.05). There was no significant association in any polymorphism (p ≥ 0.05). We also examined the association between the five variants and disease progression taking into account all possible comparisons  in cases, diseased controls and healthy controls, and the respective ORparisons .p-value = 0.037). Estimated glomerular filtration rate (eGFR) was also differed statistically significant between healthy controls, diseased controls and diabetic nephropathy cases .In addition, we examined the correlation between individual genotypes and clinical features . Only rsThe literature review identified 240 titles in PubMed that met the inclusion criteria. When an article provided data for different populations, each population was considered as a different study. SERPINE1 gene, we selected five tag SNPs for genotyping in a cohort of Greek origin and we also performed a meta-analysis that included all available genetic data regarding genetic variants of the aforementioned gene.In an effort to provide the most comprehensive overview assessing for genetic variation in SERPINE1 gene polymorphisms have been studied in various systematic reviews and meta-analyses regarding atherosclerotic diseases [diseases , risk ofdiseases , stroke diseases  and diabdiseases ,38,48.SERPINE1 gene. Similarly, the overall meta-analysis of 4G>5G variant (rs1799889) as well as the subgroup analyses based on diabetes type and ethnicity did not detect any significant association, indicating no implication of SERPINE1 variants in the risk or development of the disease.The present association study did not reveal statistical significance for any genotyped tag SNP located in SERPINE1 4G/5G polymorphism and DN in Europeans patients with T1DM [SERPINE1 and DN [In agreement to the findings of the present genetic association study and meta-analysis, De Cosmo et al. (1999) found no association between ith T1DM . Similar1 and DN . SERPINE1 4G/4G is an independent factor for the development of DN in Chinese patients with T2DM and also exhibits a synergistic effect with the allele D of ACE gene on development of DN [SERPINE1 4G allele might be risk allele for DN susceptibility in the Chinese population [In contrast to our findings, Dastgheib et al. (2020) found that the PAI-1 4G5G polymorphism was associated with increased risk of DN and diabetic retinopathy (DR) risk . Furthernt of DN . In addint of DN . Last bupulation .SERPINE1 is influenced by the ethnicity and/or type of diabetes, as many studies conducted in Asians with T2DM detected significant association between genetic variants of SERPINE1 and DN, whereas studies performed in Caucasians with T1DM did not detect any significant association between SERPINE1 and DN. Our suggestion is in agreement with the conclusion of a recent meta-analysis of SERPINE1 rs1799889 variant, in which Chen et al. (2021) concluded that 4G polymorphism could constitute a genetic synergistic factor in overall DM and DN populations, positively for individuals with Asian descent [Taking into account the discrepancy in findings derived from different study, it could be suggested that genetic variability in  descent . It is n descent .G is a model-free approach, so it takes advantage of the full genotype distribution and provides a straightforward interpretation of genetic associations. On the other hand, the present study has certain limitations. The sample size of the present case-control study is relatively small raising the possibility of false positive and false negative results, an inherent though limitation of the majority of genetic association studies. Moreover, in the present meta-analysis only studies written in English language were included.Our study design has several strengths out of which clear case definition is one of them. Patients without persistent proteinuria were not considered as cases with DN, since not persistent proteinuria could be reversible and could lead to underestimation of the genetic effect. In addition, we included in the study, healthy controls without diabetes in order to discriminate any variant correlated with diabetes mellitus but not with DN per se. Moreover, the use of ORSERPINE1 gene polymorphisms in the context of a case-control study without statistical significant results for any variant. In an effort to further examine the role of SERPINE1 gene variants in the risk of DN, we performed a systematic review and meta-analysis where 4G/5G polymorphism was also not revealed statistical significant. Further studies remain to verify the results of the present meta-analysis.In conclusion, we investigated the role of


I2 statistic and Egger’s regression test, respectively. Results: Eleven articles consisting of two case-control studies, eight cross-sectional studies, and one prospective cohort study were included in the meta-analysis. Of those, three studies with 92,075 subjects were included in the analysis for NAFLD incidence, eight studies with 9558 subjects for NAFLD prevalence, and five with 4303 subjects were used for the analysis of liver fibrosis. There was no association between coffee consumption and NAFLD incidence  or NAFLD prevalence . The meta-analysis showed coffee consumption to be significantly associated with a 35% decreased odds of significant liver fibrosis . There was no heterogeneity  and no evidence of publication bias (p = 0.134). Conclusion: This meta-analysis supports the protective role of coffee consumption on significant liver fibrosis in patients with NAFLD. However, the threshold of coffee consumption to achieve hepatoprotective effects needs to be established in prospective trials.Background and aim: Non-alcoholic fatty liver disease (NAFLD) is the most common cause of chronic liver disease worldwide. Given the anti-fibrotic and antioxidant properties of coffee, this systematic review and meta-analysis aims to provide updated results on the impact of coffee consumption on NAFLD incidence, prevalence, and risk of significant liver fibrosis. Methods: We conducted a comprehensive search in MEDLINE (OvidSP) and Scopus from January 2010 through January 2021. Relative risks for the highest versus the lowest level of coffee consumption were pooled using random-effects models. Heterogeneity and publication bias were evaluated using the Higgins’  Non-alcoholic fatty liver disease (NAFLD) is the most common cause of chronic liver disease in North America and is one of the main health complications seen around the world. Recently, a change in its nomenclature to Metabolic-Associated Fatty Liver Disease (MAFLD) ,2 has beThe estimated prevalence of NAFLD is 25–30% in the adult population . NAFLD cGenetic and lifestyle factors play an important role in the pathogenesis of NAFLD. Thus, current management focuses mainly on lifestyle adaptations to reduce steatosis and risk of fibrosis development. Numerous studies have shown beneficial effects of coffee consumption in various diseases including diabetes mellitus type 2, cardiovascular disease, HCC, and NAFLD ,10,11. TNevertheless, the effects of coffee consumption on NAFLD risk have been conflicting. Therefore, this systematic review and meta-analysis aims to provide updated results on the impact of coffee consumption on NAFLD incidence, prevalence, and liver fibrosis risk.We conducted a comprehensive search for titles, abstracts, and keywords in MEDLINE (OvidSP) and Scopus using the search term: (coffee OR caffeine) combined with  from January 2010 through January 2021. Reference lists of included manuscripts and published meta-analyses were also checked manually. The search was limited to human studies written in English.The inclusion criteria were as follows: (a) case-control, cross-sectional, or cohort studies; (b) published as original studies to evaluate the impact of coffee consumption on NAFLD incidence, prevalence, and risk of significant liver fibrosis; (c) including relative risks (RRs), odds ratios (ORs), or hazard ratios (HRs) with 95% confidence intervals (95% CIs) or adequate data to calculate them; (d) NAFLD diagnosis by serum biomarkers of liver injury or imaging studies , elastography, or liver histological evaluation.We excluded studies without sufficient statistics or adequate data for risk estimates, lack of biochemical tests and imaging to evaluate liver injury, no description of coffee intake measurement , use of coffee derivatives, no discrimination between coffee or caffeine as the main exposure, or if NAFLD incidence, prevalence, or significant liver fibrosis was not the main outcome of interest. Studies evaluating the association between coffee consumption and NAFLD in the general population or in patients with NAFLD are included. We consider significant liver fibrosis as a histological score greater than or equal to F2 according METAVIR or elastography >8 kPa. All these studies assessed liver fibrosis in a single time-point and therefore, we present the results as liver fibrosis associated with NAFLD but not the fibrosis progression in NAFLD, as assessing progression requires more than one-time measurements.For each study, the author names and year of publication, data on study design, population, NAFLD diagnosis, coffee consumption assessment, steatosis and fibrosis severity classification, RRs/HRs/ORs estimates for extreme categories of coffee intake and corresponding 95% CIs, the adjustments for confounding variables, and main findings were extracted. We did not communicate with the study authors regarding missing information. The methodological quality of each article was assessed using the Newcastle−Ottawa Scale (NOS) , which aIn this meta-analysis, ORs of the case-control and the cross-sectional studies as well as HRs of the cohort study were used as an estimate of RRs. RRs for the highest versus the lowest level of coffee consumption were pooled using random-effects models. If more than one RR estimate was reported, the one most comprehensively adjusted for potential confounding factors was used. For studies reporting only the number of cases and controls for each category of coffee consumption, crude RRs and the corresponding 95% CIs were computed from these frequencies. Studies with risk estimates of two categories  were also integrated. I2 statistic with values ≤25%, ≤50%, ≤75%, and >75% representing no, small, moderate, and significant heterogeneity, respectively [Egger < 0.1. All analyses were performed using Stata 16.0  and Review Manager (Rev-Man) version 5.4 with p-values < 0.05 considered as statistically significant.Heterogeneity was evaluated using the Higgins’ ectively . We alsoectively  with theThe initial search strategy yielded 303 potentially relevant articles after the exclusion of duplicates. Two hundred and sixty-seven articles were excluded after screening titles and abstracts, leaving the remaining 36 articles for a full-text article review. According to inclusion and exclusion criteria, 11 articles fulfilled the eligibility criteria and, therefore, were included in the data analysis ,27,28,29Data on coffee consumption was collected by interviewer-administrated questionnaire in three studies ,25,29. Tn = 147) was prospectively followed for 7 years, no association was found between coffee intake and NAFLD onset assessed by liver ultrasound . Although poorer lifestyles were noticed in subjects with high coffee intake (three cups/day), controlling for lifestyle factors  did not reveal any association with NAFLD onset [Association between coffee consumption and NAFLD incidence, prevalence, and liver fibrosis risk was investigated in previous observational studies. In a cross-sectional study of 347 subjects, an inverse association was reported between coffee consumption (3 cups/day) and degree of liver fibrosis, but there was no association with NAFLD prevalence. Similarly, when a subgroup of subjects without fatty liver at baseline (LD onset . No assoLD onset .p = 0.02) [A recent cross-sectional study in 555 NAFLD patients found no association between coffee drinking and severe hepatic steatosis, defined as CAP 280 dB/m. Coffee consumption was associated with a lower risk of advanced fibrosis (defined as liver stiffness measurement (LSM) ≥10 kPa) after adjusting for confounders  . Increased intake of coffee, rather than total caffeine intake from other sources, was protective against the progression of fibrosis in patients with NASH. In a cross-sectional study of biopsy-proven NASH patients and ultrasound-negative controls, coffee intake was higher in controls and patients with NASH stage 0–1 compared to those with stage 2–4 fibrosis, suggesting the beneficial effects of coffee in preventing fibrosis development . The metA dose-dependent protective effect of coffee on the prevalence  and sevep = 0.48; I2 = 83%, p = 0.003) but no evidence of publication bias (p = 0.674). In a sensitivity analysis that excluded the study by Chung et al. [I2 = 0%, p = 0.96) was associated with a significant reduction in the incidence of NAFLD . Due to the small number of studies, we were unable to conduct subgroup analyses to determine the sources of heterogeneity. However, NAFLD has a prolonged natural course, and variable follow-up times between studies ranging from 2.8 years in the study by Chung et al. [During the three to seven years’ follow up, no association was seen between coffee consumption and NAFLD incidence in subjects without NAFLD at baseline  or publication bias (p = 0.699). To check the effect of individual studies on the pooled RR, a sensitivity analysis was performed with exclusion of one trial [p = 0.04). There was no significant change in RR or 2I values with removal of other studies. Subgroup analyses according to the study design, questionnaire administration, study quality, and energy intake adjustment revealed that the association was significant in studies with lower quality, without adjustment for total energy intake, or when the self-reported questionnaire was applied . Egger’s linear regression test was performed and showed no publication bias (p = 0.134). Sensitivity analysis by excluding one study at a time revealed high stability of the results and no single influential study on the pooled RR, which varied from 0.59 (95% CI 0.43–0.80) when the study from Bambha et al. [p = 0.0004) and the general population .Meta-analysis with a randomized-effect model including five studies showed that coffee consumption was significantly associated with 35% decreased odds of liver fibrosis  with either the prevalence or the risk of liver fibrosis related to NAFLD. However, when stratified by the type of caffeine intake, regular coffee caffeine intake (mg/day) significantly decreased the risk of liver fibrosis or development of NAFLD . Lower rSome limitations should be noted in interpreting the results of the current meta-analysis. First, this analysis was limited to the English language and exclusion of manuscripts in other languages or unpublished studies might contribute to publication bias. The majority of the studies in this area were cross-sectional studies showing an association but not able to determine causality and limiting the capability to evaluate chronological associations. Reverse causality is a common concern in cross-sectional studies as the predictor and outcome are assessed at the same time. The risk of bias and the quality of the studies were assessed using the NOS, a commonly used tool for observational studies. Although a conventional sampling method was applied in most of the studies, comparison between responders and non-responders was not reported, contributing to self-selection bias. Observational studies are limited by the inability to account for confounding factors, including nutritional and lifestyle behaviors . Several comorbidities affecting the severity of NAFLD such as diabetes and obesity differ between coffee drinkers and non-drinkers, and a lack of adjustment can lead to bias in the assessment of the effect of the exposure. Daily energy intake may be higher in subjects who regularly drink coffee with additives such as sugar and cream , and witOne of the main challenges in these studies is to properly assess an individual patient’s coffee consumption. Coffee intake was evaluated using a questionnaire at a single time-point and therefore the impact of changes in habitual coffee intake was not determined. The validity of the questionnaire and blinded interview are also important considerations. Retrospective self-reported questionnaires infer the risk of recall or reporting bias . IllustrPotential beneficial components of regular coffee need to be determined in prospective studies as coffee contains several compounds including caffeine, polyphenol (chlorogenic acid), and diterpenes (cafestol and kahweol) . ExperimThis meta-analysis confirms the protective role of regular coffee consumption on the risk of significant liver fibrosis. No association was observed between coffee consumption and NAFLD incidence or prevalence. In the absence of approved drug therapy, coffee should be recommended in conjunction with lifestyle changes. The threshold of coffee consumption to achieve its protective effect needs to be established in prospective trials. It is still unclear which element of coffee confers benefit. Prospective studies with a comprehensive evaluation of coffee intake over time will be important to clarify the amount, type, preparation method, and pattern of coffee consumption that has the most benefit in liver disease. The impact of possible confounding factors such as enhanced nutritional status or physical activity cannot be excluded.


Background and Objective: An increasing number of epidemiological original studies suggested that long-term exposure to particulate matter (PM2.5 and PM10) could be associated with the risk of myocardial infarction (MI), but the results were inconsistent. We aimed to synthesized available cohort studies to identify the association between ambient air pollution (PM2.5 and PM10) and MI risk by a meta-analysis.Methods: PubMed and Embase were searched through September 2019 to identify studies that met predetermined inclusion criteria. Reference lists from retrieved articles were also reviewed. A random-effects model was used to calculate the pooled relative risk (RR) and 95% confidence intervals (CI).Results: Twenty-seven cohort studies involving 6,764,987 participants and 94,540 patients with MI were included in this systematic review. The pooled results showed that higher levels of ambient air pollution (PM2.5 and PM10) exposure were significantly associated with the risk of MI. The pooled relative risk (RR) for each 10-μg/m3 increment in PM2.5 and PM10 were 1.18 (95% CI: 1.11–1.26), and 1.03 (95% CI: 1.00–1.05), respectively. Exclusion of any single study did not materially alter the combined risk estimate.Conclusions: Integrated evidence from cohort studies supports the hypothesis that long-term exposure to PM2.5 and PM10 is a risk factor for MI. The incidence and prevalence of cardiovascular diseases have increased in recent decades, and cardiovascular disease has become one of the main causes of death among adults –3. MyocaAir pollution, especially particulate matter (PM), has been increasingly investigated as an environmental risk factor for MI morbidity and mortality recently. However, studies have had modest sample sizes and reported inconclusive results –13. ThesTaking into consideration the inconsistent conclusions and the limitations of existing epidemiological studies and the flaws of previous meta-analyses, we therefore performed a systematic review and meta-analysis of cohort epidemiological studies to examine the potential associations between air pollution exposure and the risk of MI. Given the heavy economic and health burden of curing MI, the results of our study may provide additional practical and valuable clues for the prevention of MI.Ethical approval is not required for this systematic review.10”or “PM2.5” or “air quality” in combination with “myocardial infarction” or “heart attack” or “acute coronary syndrome” or “cardiovascular disease” or “heart disease.” The language was restricted to English. Additionally, reference lists of the retrieved original articles and relevant review articles were also scrutinized to identify further pertinent studies.We conducted this meta-analysis in accordance with the Preferred Reporting Items for Systematic Reviews and Meta-Analysis (PRISMA)  and the Studies meeting the following criteria were included in the meta-analysis: (1) the study design was cohort; (2) the exposure of interest was ambient air pollution; the endpoint of interest was the incidence of MI; and (3) the relative risk (RR) and the corresponding 95% confidence interval (CI) of MI relating to ambient air pollution were reported or could be calculated from the data provided. Animal studies, clinical trials, reviews, letters, and commentaries were excluded. Only studies with detailed information on both ambient air pollution and the incidence of MI was included.Two investigators (WF and SC) extracted the following information from the studies: first author, publication year, country, study period, age, number of cases, size of cohort, and time windows of exposure. Discrepancies were resolved by discussion with a third investigator (SC).The Newcastle-Ottawa Scale, used to evaluate the qualities of cohort studies , is a niWe used RR to measure the association between ambient air pollution and the risk of MI and the random effects model was used to calculate an overall pooled RR for the main analysis.Q statistic with a significance level at P < 0.10 and I2 statistic were used to test heterogeneity. The I2 statistic measures the percentage of total variation across studies due to heterogeneity rather than chance. It was calculated according to the formula by Higgins , MI mortality (twelve studies), and only one study MI hospital, and two studies included MI incidence and mortality. Twenty studies were published after 2010, and seven studies before 2010. The quality assessment scores ranged from 6 to 9, with an average score of seven points, representing satisfactory quality of the studies.The main characteristics of the 27 studies on MI and long-term PM exposure in our meta-analysis were summarized in 2.5 and PM10 with MI.We employed meta-analysis to assess the association of PM2.5 and PM10) and the risk of MI were shown in 2.5 exposure on the risk of MI and the pooled estimates suggested a positive relationship. Results from standardized data showed that a 10 mg/m3 increase in PM2.5 exposure was positively associated with the risk of MI , and there was a moderate heterogeneity . Subgroup analyses by state, and publication year (before 2010 vs. after 2010) showed no statistically significant difference in results , and the heterogeneity is . We conducted subgroup analyses by state, and publication year (before 2010 vs. after 2010). In general, these subgroup analyses showed no statistically significant difference in results.Fourteen studies investigated the association of PMsk of MI . The poo2.5 and PM10) and MI risk, to examine the influence of various exclusions on the combined RR and assess the robustness of all results. The pooled RR did not materially change, for PM2.5 and PM10, the both overall combined RR did not materially change, with a range from 1.17 (95% CI: 1.10–1.24) to 1.19 (95% CI: 1.11–1.26), and 1.02 (95% CI: 1.00–1.04) to 1.03 (95% CI: 1.01–1.05), respectively.Sensitivity analyses was conducted to find potential sources of heterogeneity in the association between air pollution . We used the trim-and-fill method to evaluate the impact of any potential publication bias, and the results showed that one and five potentially missing studies would be needed to obtain funnel plot symmetry for the association of MI and air pollution , respectively. After using the trim-and-fill method, both the corrected RR for PM 2.5 was 1.18 , for PM10 was 1.03 , which suggested the both pooled RRs were not substantially changed by the correction for potential publication bias.Visual inspection of the funnel plot showed significant asymmetry (PM2.5 and PM10) and MI. The pooled analysis, which included 27 cohort studies with more than 6.5 million people, showed an inverse association between exposures to air pollutants (PM2.5 and PM10) and the risk of MI, which was consistent with previous reviews and meta-analyses , which was inconsistent with previous studies. The possible reasons for the differences include inconsistency of study designs and potentially selective reporting of the results for pollutants. However, it is suggested that the Asia region should pay attention to the relationship between air pollution and MI, and more work on air pollution and epidemiology remains to be done. We conducted a subgroup analysis by publication year but found the pooled result of studies before 2010 was not significantly different from that after 2010.Considering people's different diets and lifestyles and the prevalence of MI in different regions, we also conducted subgroup analysis by region, and statistically significant differences across subgroups were found except for PMThere are several strengths in this meta-analysis. First, all the studies in our analysis were cohort studies, which is considered as stronger measure for demonstrating causation and identification of risk factors than other observational study designs . Second,2.5 and PM10. A compounding effect of all these air pollutants should be examined and quantified on increasing the risk of MI.Some limitations in the present meta-analysis should be of concern. First of all, the heterogeneity of the studies included was significant and existed through the whole analysis. But we explored the potential heterogeneity resources by subgroup analysis and sensitivity analysis. Second, there are relatively few studies on the mortality and hospitalization of MI, and there is a lack of data required for meta-analysis, so subgroup analysis is not conducted. Third, other air pollutants may have interaction with PM2.5 and PM10 as a significant risk factors of MI. In light of the heavy economic burden of MI, the results of our study provide additional valuable clues for the prevention of MI. For future studies, more high-quality longitudinal and interventional studies are needed to explore the underlying mechanisms of the relationships between ambient air pollution and MI, especially in Asia and other middle- and low-income countries. Besides, it is recommended that relevant departments must adopt a comprehensive particulate matter control policy to promote health and reduce the burden of MI.In conclusion, our meta-analysis identified long-term exposure to PMThe original contributions presented in the study are included in the article/LZ, QZ, YZ, SC, and CL conceived and designed the study. LZ, QZ, WF, ZZ, HX, SY, JM, YZ, SC, and CL participated in the acquisition of data. LZ and QZ analyzed the data. YZ, CL, and SC gave advice on methodology. LZ drafted the manuscript. LZ, QZ, WF, ZZ, HX, SY, JM, YZ, SC, and CL revised the manuscript. All authors read and approved the final manuscript.The authors declare that the research was conducted in the absence of any commercial or financial relationships that could be construed as a potential conflict of interest.


Several recent studies have investigated the prognostic and clinicopathological significance of epithelial cadherin (E-cadherin) in pancreatic cancer; however, conclusions from these studies remain inconsistent. Therefore, we performed a meta-analysis to evaluate the effects of E-cadherin expression on the prognosis and clinicopathological characteristics of pancreatic cancer.2 statistic.Embase, PubMed, and Web of Science were searched to identify articles associated with E-cadherin and pancreatic cancer. Hazard ratios (HRs) and odds ratios (ORs) with corresponding confidence intervals (CIs) were calculated and summarized. All eligible studies were searched until May 20, 2020. Heterogeneity among studies was assessed using the Chi-square test and IOverall, 25 studies were identified, of which 12 reports with 1,032 cases concerned the prognosis of pancreatic cancer, and 22 involved the risk and clinical characteristics of pancreatic cancer. The overall results revealed that E-cadherin expression was significantly related to overall survival, gender, tumor grade, lymph node metastasis, tumor differentiation, and risk of pancreatic cancer. In the subgroup analysis, no significant heterogeneity or publication bias was observed.E-cadherin expression is strongly associated with the risk, clinical features, and prognosis of pancreatic cancer, suggesting that E-cadherin may be an effective biomarker for the clinical assessments and predicting prognosis of pancreatic cancer. Pancreatic cancer is one of the deadliest human malignancies, with a reported 5-year relative survival rate of less than 5% . AccordiEpithelial cadherin (E-cadherin) is a typical type I classical cadherin, which is expressed mainly in epithelial cells and can mediate cell-cell interactions . PublishEligible reports regarding the association of E-cadherin expression with clinicopathological features and prognosis of pancreatic cancer were systematically searched in electronic databases, including PubMed, Embase, and Web of Science, until May 20, 2020. We searched these electronic databases using the following keywords: “pancreatic cancer,” “pancreatic carcinoma,” “E-cadherin,” “cadherin,” “CDH1” and “prognosis.” Two independent investigators collected the included studies based on the inclusion and exclusion criteria. Furthermore, the references cited in included studies were screened to identify other eligible studies.Eligible studies were included according to predefined inclusion and exclusion criteria. All included studies had to meet the following inclusion criteria: (1) studies published in English; (2) case-control studies, retrospective studies, or cohort studies; (3) patients had to present detailed pathological diagnostic criteria; (4) studies had to detail sufficient data to calculate hazard ratios (HRs), odds ratios (ORs), and 95% CIs. Conversely, studies were excluded based on the following exclusion criteria: (1) reviews, comments, letters, conferences, case reports, editorials, and meta-analyses; (2) studies with duplicated data; (3) no available data.http://www.ohri.ca/programs/clinical_epidemiology/oxford.asp, accessed on June 1, 2019) was applied to comprehensively evaluate the quality of included studies that involved the relationship between the risk of pancreatic cancer and E-cadherin expression. The included studies were categorized as low (<4), medium (4–6), or high (>6) according to their quality scores , survival curve, and the number of cases and controls in different groups. The Newcastle–Ottawa Scale (NOS) table . Initially, HRs, ORs, 95% CIs, and the weight of each included study were calculated with a fixed-effect model and random-effects model to estimate the role of E-cadherin expression in risk, prognosis, and clinicopathological features of pancreatic cancer . Statist studies . If sign studies . Begg’s  studies . In addi studies . FinallyOf the 381 studies initially identified, 12 studies involving 1,032 pancreatic cancer patients analyzed the influence of E-cadherin on the prognosis of pancreatic cancer, whereas 22 studies assessed the risk and clinicopathological features of pancreatic cancer –43. HoweThe overall results indicated a significant relationship between E-cadherin expression and overall survival (OS) of pancreatic cancer . No significant heterogeneity was detected; therefore, a fixed-effects model was used to calculate the pooled HRs. The pooled HR was 1.93, indicating that decreased expression levels of E-cadherin were significantly associated with the poor OS of pancreatic cancer. To further assess the association between aberrant E-cadherin expression and OS in pancreatic cancer, subgroup analyses based on race, source of HRs, and cancer subtype were conducted. The subgroup analysis showed similar results both in Caucasian  and Asian populations . In addition, the subgroup analysis for pancreatic ductal adenocarcinoma (PDCA) revealed that low E-cadherin expression led to a poor OS in pancreatic cancer . None of the subgroup analyses detected significant heterogeneity across studies Table 3 Among the included studies, four investigated the association between E-cadherin expression and the risk of pancreatic cancer, and a significant association was observed among Caucasian and Asian populations . Therefore, decreased E-cadherin expression may lead to an increased risk of pancreatic cancer. Next, we assessed the correlation between decreased E-cadherin expression and the clinicopathological features of patients with pancreatic cancer. The results are summarized in 2 statistics, heterogeneity was not detected among included studies , lymph node metastasis  and metastasis , and differentiation  in pancreatic cancer. Based on Cochran’s Q and I studies Table 5 Sensitivity analysis was performed to examine the effects of each included study on pooled HRs and ORs. For OS in pancreatic cancer, the results of Wang et al.  significDuring EMT, the cobblestone epithelial appearance of cultured epithelial cells gradually changed to a spindle-shaped, mesenchymal morphology. This transformation is reversible, in which mesenchymal cells can convert back into an epithelial state, known as a mesenchymal–epithelial transition (MET) . EMT plaIn the present meta-analysis, the pooled results suggested that cancer tissues present a lower E-cadherin expression than normal tissues or adjacent tissues , 28, 31,Tissue samples from different subtypes of pancreatic tumors were included in the current meta-analysis. Therefore, subgroup analysis based on cancer type was performed to investigate the association between E-cadherin expression and PDCA, which is the most frequent type of pancreatic cancer. A significant correlation was detected between E-cadherin downregulation and the prognosis and clinical progress of PDCA. Furthermore, we observed a more obvious association between E-cadherin downregulation and poor PDCA differentiation. In addition, according to the pooled ORs, E-cadherin downregulation promoted the metastasis of PDCA cells. These observations were in line with the perspective that E-cadherin is a tumor suppressor molecule in cancer development , which fThere is growing evidence that the tumor microenvironment (TME) plays an important role in the progression and prognosis of pancreatic cancer. The TME consists of the extracellular matrix (ECM), aberrant immune cells, pancreatic stellate cells, and abnormal tumor vascularization . It has Based on findings of the current meta-analysis, E-cadherin expression was significantly decreased in pancreatic cancer tissue samples and presented a close relationship with poor prognosis and malignant progression of the tumor patients. In the analysis of pancreatic cancer differentiation, we detected heterogeneity among studies of Asian populations; accordingly, subgroup analysis based on cancer type was further performed in PDCA. No heterogeneity was observed in the analysis of PDCA. Additionally, the sensitivity analysis results revealed that no individual study significantly affected the stability of the pooled HRs and ORs.Although heterogeneity and publication bias were rarely observed, some limitations persist in the present study. First, we extracted HRs from the survival curve when eligible studies did not directly provide HRs, and the extracted HRs might minimally differ from the real data. Second, the sample number in some subgroup analyses was small. Third, although we included all eligible literature in English, several studies in other languages were not included, which might have excluded some potential data.In conclusion, this meta-analysis demonstrated that decreased E-cadherin expression was positively correlated with advanced clinical features and poor OS in patients with pancreatic cancer, suggesting that it might be a potential biomarker for clinical assessments and predicting the prognosis of pancreatic cancer.The original contributions presented in the study are included in the article/All authors listed have made a substantial, direct, and intellectual contribution to the work and approved it for publication.The authors declare that the research was conducted in the absence of any commercial or financial relationships that could be construed as a potential conflict of interest.


This study investigates potential uncontrolled confounding in meta-analyses of the association of being overweight with all-cause mortality. Just as individual nonrandomized studies can be biased because of uncontrolled confounding,3 so can meta-analyses.4 In both meta-analyses, many of the included studies did not control for probable confounders . First, we calculated an E-value representing the minimum strengths of associations on the risk ratio (RR) scale that uncontrolled confounders would need to jointly have with being overweight and with mortality across all studies in each meta-analysis to shift the meta-analytic estimate or its 95% CI to the null , this may suggest that most studies have meaningfully strong HRs, albeit prior to considering uncontrolled confounding.5 We then assessed how strong uncontrolled confounding would have to be to reduce the percentage of meaningfully strong HRs to 15%.5 For code and data for reproducibility, see eMethods in the Second, we estimated the percentage of studies with meaningfully strong HRs , initially without considering potential uncontrolled confounding.P < .001; heterogeneity τ̂ = 0.13) for Flegal et al1 and 1.10  for GBMC.2 These analyses comprised, respectively, 140 and 186 estimates from prospective nonrandomized cohorts.Before considering uncontrolled confounding, our reanalysis obtained pooled HRs of 0.93  of studies in GBMC2 but 9%  of studies in in Flegal et al1 would have meaningfully strong detrimental HRs . Again, these percentages refer to the heterogeneous distribution of associations that underlie each meta-analysis. However, small uncontrolled confounding associations  would suffice for both meta-analyses to bring the percentages of causal effects with HRs below 0.9  or above 1.1 (for GBMC2) to the same level (15%) in the 2 meta-analyses  as 40%  in Flegal et alanalyses .This reanalysis of meta-analyses found that for 2 meta-analyses, uncontrolled confounding associations in each study with RRs of 1.25 to 1.43 could potentially shift the point estimate or CIs to the null or substantially decrease the percentage of meaningfully strong effect sizes. These small degrees of uncontrolled confounding seem plausible in the context of studies’ limited control of confounding by physical, social, behavioral, and psychological factors. These sensitivity analyses suggest that neither meta-analysis provided robust evidence for protective or detrimental potential effects of being overweight on mortality.6 We found that uncontrolled confounding may temper interpretation of meta-analyses. The differing results of Flegal et al1 and GBMC2 may reflect heterogeneous effects and differing inclusion criteria, along with differing uncontrolled confounding. Establishing potentially modest effects of being overweight on mortality would require improved study designs for primary studies and meta-analyses alike.4Our analysis has several limitations. First, we considered bias of constant severity across studies. Second, thresholds defining associations with meaningfully large HRs  are somewhat arbitrary and should be informed by scientific context.


Dear Editor,We appreciate all of the reviewers for their thought-provoking comments about our recent meta-analysis published in the journal . Our ansFirstly, although the heterogeneity of total blood loss is high, we use the random effects model that allows reliable pooled results; this heterogeneity might be attributed to different surgical procedures. At present, there is no conclusion about the data-base to be searched. Many meta-analyses also searched only two or three databases –11. The It is well known that the prevalence of knee osteoarthritis is much higher in female. Meta-regression analysis for gender as an item does not make much sense. We do not think it is appropriate to conclude that female might benefit more than male on blood management from tranexamic acid (TXA) by dividing them into three subgroups based on gender. From their results, it can only be concluded that no matter what the proportion of female is, TXA can benefit. Moreover, gender difference could affect total blood loss, with greater amount in men compared with women –15.Secondly, we agree with the hypothesis that intraoperative TXA had a short-time effect, but it might benefit patients for a relatively long time. This is consistent with the results of our meta-analysis, which is also in agreement with previous studies , 17.P = 0.247) indicated a low risk of publication bias. However, as this meta-analysis did not include enough studies, the reliability of these assessments was not very strong. Strictly speaking, publication bias in this study is not necessary.Finally, most of the studies included in this meta-analysis are cohort studies, which cannot be as consistent as randomized controlled trials (RCTs). And we choose random-effect model to make the results tend to be conservative. Moreover, the results were consistent even with the fixed-effect model Figs.  and 2. T


To investigate the impact of large choroidal vessels (LCV) on Choriocapillaris (CC) flow deficit (FD) analyses with swept-source optical coherence tomography angiography (SS-OCTA)Prospective, cross-sectional study.Macular 6x6mm SS-OCTA scans were obtained from intermediate age-related macular degeneration (iAMD) and healthy eyes. Images were captured and processed according to most common standards and analyzed for percentage of flow-deficits (FD%) within four 1x1mm squares at the corners of each image. Choroidal thickness (CT), iris color and refraction error were considered as potential influential factors for LCV visibility. A linear mixed model and logistic regression models were calculated for statistical evaluation.Sixty-nine iAMD and 49 age-matched healthy eyes were enrolled. LCV were visible in at least one sector in 52% of iAMD and 47% of healthy eyes. Within the iAMD group FD% were significantly lower in areas containing LCV (p = 0.0029). Increasing CT resulted in an odds ratio decrease of LCV . Below a CT value of ≤118μm LCV could be expected with a sensitivity of 86% and a specificity of 85%.LCV can significantly affect CC FD analyses of SS-OCTA images. Their visibility is negatively associated with CT. The impact of LCV should be taken into account when performing CC FD assessments, especially in patients where reduced CT is to be expected and inclusion of affected areas should be considered carefully. The choriocapillaris (CC) marks the capillary endpoint of the choroid and is located right below the Bruch´s Membrane (BM)/retinal pigment epithelium (RPE) complex. The choroid is subdivided into three vascular layers, the Haller layer, the Sattler layer and the CC, with vessel diameters becoming progressively smaller as they get closer to BM. The anatomical transition between choroidal vascular layers is fluent and without consensus definition at which diameter vessels are considered small, medium or large sized . With veThis cross-sectional study was performed in accordance with the tenets of the Declaration of Helsinki and was approved by the Ethics Committee of the Medical University of Vienna. Written informed consent was obtained from all participants prior to enrolment.Consecutive patients suffering from intermediate AMD (iAMD)   were incTM motion correction was used during imaging.6x6mm swept-source optical coherence tomography angiography (SS-OCTA) images with a resolution of 1024x1024 pixels, centered on the fovea were acquired using a Zeiss Plex Elite 9000 . This instrument uses a 100-kHz light source with a central wavelength of 1060nm, a full-width at half maximum axial resolution of ~5μm in tissue, a lateral resolution at the retinal surface estimated at ~14μm and an A-scan depth of 3.0mm in tissue. Each 6x6mm scan consists of 500 twice repeated A-scans per B-scan. To minimize subject motion artifacts, FastTracImage acquisition and post-processing parameters were based on most commonly used standards in literature to obtain comparable results , 9, 13. To compensate for image artifacts arising from light attenuation, a signal compensation method using structural information from the same slab, was performed . Images Statistical analyses were conducted with SAS  and R  p-values < 0.05 were considered statistically significant. Quantitative demographic and clinical variables are reported as mean values ± standard deviation (SD) and group comparisons were done by two-sample t-tests for unequal variances. For qualitative variables, absolute frequencies are reported and group comparisons were done by Chi-squared tests.To investigate, whether LCV (1 measurement per quadrant) had an effect on FD analyses, a mixed model was calculated with FD% values used as the dependent variable. Presence of LCV, topographic orientation , group  and the interactions of LCV and orientation, LCV and group, orientation and group were defined as independent variables. Patient was included as a random effect to account for the repeated measurements per eye. As the F-test of the interactions between LCV and orientation, and between orientation and group had a p-value greater than the significance level of 0.05 (results not shown), these interaction-terms were removed from the model. The comparison between the iAMD and the control group as well as the comparison between sectors containing LCV vs sectors without LCV were done with t-tests derived from the mixed model. The 10μm thick slab starting 31μm below the RPE-fit line, including signal compensation, a Phansalkar radius of 15 pixels and exclusion of FDs smaller than the average ICD of 24μm was chosen as our main setting. For all other settings, the analysis was done in an exploratory way, therefore we adjusted within each endpoint for the number of group comparisons  but not for the number of settings.To investigate the effect of group , CT and orientation  on visibility of LCV, univariate logistic regression models were calculated with patient set as a random factor. If a univariate model revealed a p-value < 0.05 the independent variable was included in a multivariable logistic regression model with LCV visibility set as the dependent variable and patient as a random factor to account for multiple measurements within the same eye. The effect of eye color, cCT and spherical equivalent on LCV visibility was analyzed by univariate logistic regression models, whereby the dependent variable was defined as “presence of LCV in at least one sector”. Due to the exploratory character of these analyses we did not adjust for multiple testing. Hence, the interpretation of the results is descriptive.To determine a cutoff, that differentiates between perceptible LCV (LCV = 1) and imperceptible LCV (LCV = 0), a univariable logistic regression model was calculated with CT (measured in 4 areas) as independent variable. The best cut-off was selected by the maximal Youden-Index. Based on this cut-off, sensitivity and specificity values were calculated. To consider the repeated measurements per patient, the 95% confidence limits for sensitivity and specificity were calculated by adjusting the confidence limits using a ratio estimator . As thesSixty-nine iAMD and 49 age-matched healthy eyes were enrolled, resulting in a total of 118 eyes from 118 patients. The mean age was 73±6.1 years  in the iAMD group and 72±8.6 years  in the healthy control group. The overall demographic and clinical characteristics of the two groups are shown in the provided In our mixed model the presence of LCV, group (iAMD and control group), and the interaction between those two parameters (p = 0.0174) had a statistically significant impact on FD% analyses. If LCV were visible, mean FD% values were lower within the iAMD group ). Within the control group, this difference was not statistically significant  . The iAMThe other OCTA-slab and post-processing settings yielded similar results with exception of the 10μm thick slab starting 31μm below the RPE-fit line using a Phansalkar radius of 4, which was not able to differentiate between the iAMD and the control group, in either signal compensated, or uncompensated images. However, even with these settings, areas containing LCV yielded significantly lower FD% values within the iAMD group .Furthermore, the mixed model revealed a statistically significant effect of topographic orientation  on FD% values , with nasal sectors showing lower FD% values across both groups .The frequency of LCV did not differ significantly between the two groups (p = 0.99) with LCV visible in 26% and 27% of all sectors within the iAMD and control group, respectively . LCV werWith increasing CT, the odds for visibility of LCV decreased  . In nasaCalculating a multivariable logistic regression model, the interaction between the effects CT and orientation was also statistically significant (p = 0.0343). In temporal areas, the OR for CT was 0.958 . In nasal areas, the OR for CT was 0.931  which indicates that the effect of CT was greater within nasal areas compared to temporal areas.The effect of central choroidal thickness (cCT) on presence of LCV in at least one sector was statistically significant . With increasing cCT, the odds for visibility of LCV decreased  .Eye color  and spherical equivalent  did not affect visibility of LCV.The CT cutoff analysis showed that below a CT value of ≤118μm LCV could be expected with a sensitivity of 86%  and a specificity of 85% .In this prospective cross-sectional study, the effect of LCV on CC FD assessments was investigated. FD% values were significantly lower in areas containing LCV in the iAMD group, resulting in seemingly ‘healthier’ sectors. LCV were visible in at least one sector in 52% and 47% of iAMD and healthy eyes respectively. A negative association between the visibility of LCV and CT was found in all groups.As the lateral resolution of current OCTA devices of approximately 14μm is insufficient to resolve flow information of individual capillaries within the CC, presently used strategies inverted the analyzing process and focused on quantification of areas with decreased flow signals rather than individual vessels, so called FDs . HoweverWhen analyzing the CC, a number of factors has to be taken into consideration to obtain reliable results, ranging from anatomical properties of the layer itself like CC- and choroidal thickness, close proximity to the highly light scattering RPE/BM complex, or imaging and post-processing settings that can influence the results , 17, 21.The CC has a thickness of approximately 10μm , yet resTo compensate for light attenuation caused by the RPE/BM-complex a signal compensation algorithm based on structural information of OCT scans was developed . To testCT is negatively correlated with age  and variDue to a lack of consensus on image acquisition and post-processing operations, different strategies have been described in literature . One of However, whereas the standard 15-pixel radius allowed for appropriate distinction between healthy and diseased subjects in this study as well as in previous studies, the adjusted 4-pixel radius resulted in a reduced difference between the two groups. This effect was most pronounced when analyzing the 10μm thick slabs starting at 31μm below the RPE-fit line where no statistically significant difference between the two groups could be found anymore when applying the 4-pixel radius even though CC density was shown to decrease with AMD progression histologically [The image analysis approach chosen for this paper analyzes CC FDs at the corners of each image . This meAnother approach that might offer a solution is the averaging of multiple en-face images which can improve the visualization of capillaries and change the granular pattern of single CC OCTA-scans to a distinguishable vascular meshwork . This apAlthough the detection of LCV in CC en-face projections was performed by two independent retinal experts, their grading was not performed based on a vessel thickness threshold in μm but on appearance only. This is partially attributed to the fact that there is no clear consensus at what point vessels within the choroid are to be considered small, large or medium sized . To compThis study addresses alterations that occur in CC image analyses in the presence of LCV within the CC slab, offers an explanation for their visibility and a CT-threshold at which their perceptibility is very likely. Future studies are needed to establish compensation strategies, or vessel exclusion algorithms similar to approaches used for the superficial plexus.In conclusion, visible LCV can significantly affect the quantification of CC FD analyses of SS-OCTA images by masking possible FDs. Visibility of LCV is negatively associated with CT, making peripheral areas where CT is usually lower particularly susceptible for this disruptive element. The impact of LCV should be taken into account when performing CC FD assessments, especially in elderly patients, or patients where reduced CT is to be expected and inclusion of affected areas should be considered carefully.


Recently, some studies have suggested that the association of apurinic/apyrimidinic endonuclease 1 (APE1) gene polymorphism with prostate cancer (PCa) risk, but there are still some controversies. Hence, we elaborated the relationship between APE1 rs1760944 and rs1130409 gene and PCa risk through systematic literature review and meta-analysis.As of March 2020, EMBASE, PubMed, the Cochrane Library, Science Direct/Elsevier, MEDLINE and CNKI were used for systematic literature retrieval to investigate the correlation between APE1 rs1760944 and rs1130409 gene polymorphism with PCa risk. Meta-analysis was performed using Review Manager and Stata software.P < .05). The combined odds ratios and 95% confidence intervals were as follows: Additive model ; Codominant model ; Dominant model ; Recessive model ; Allele model . There also have significant associations between APE1 rs1130409 polymorphisms and PCa in all genetic models (P < .05). The combined odds ratios and 95% confidence intervals were as follows: Additive model ; Codominant model ; Dominant model ; Recessive model ; Allele model .Seven studies were distinguished, consists of 1769 cases of PCa patients and 2237 normal controls. Our results illustrated that there are significant correlation between the APE1 rs1760944 gene polymorphism and PCa in all genetic models (This study suggests that APE1 rs1760944 polymorphisms might be a protective factor of PCa, and APE1 rs1130409 is suggested to be a risk factor of PCa. APE1 rs1760944 and rs1130409 polymorphisms may be used in the risk assessment of PCa. Currently, much evidence suggests that complex interactions between genetic and environmental factors are involved in the development of PCa.–8 A genome-wide association study (GWAS) found that at least 35 loci were associated with PCa.,7 About 42% of PCa risk can be explained by genetic factors. It is of great value and significance to test these risk alleles in the population.Prostate cancer (PCa) is one of the most common tumors in men, causing more than 250,000 deaths every year. DNA repair gene polymorphism may cause this variation. Apurinic/apyrimidinic endonuclease 1 (APE1) in human is located on chromosome 14q11.2, which consists of five exons and spans about 2.5 to 3 kb of DNA.,11 In addition to its role in DNA repair, APE1 is also known as a transcriptional coactivator of many transcription factors, such as AP-1 HIF-1 α, p53 and NF-κB, which are involved in cancer promotion and progression. APE1 is a promising target for pharmacological treatment of some cancer types. To date, epidemiological studies have shown that single nucleotide polymorphisms (SNPs) in APE1 may confer an individual's susceptibility to PCa.,15 Some related studies have reported the correlation between the SNP variation of APE1 rs1760944 and rs1130409 and PCa risk, but the results on this issue are contradictory. Some studies reported that this polymorphism may increase the risk of PCa, but other studies did not.–22 Therefore, we systematically reviewed the available literature and performed a meta-analysis to evaluate the the relationship between APE1 rs1760944 and rs1130409 gene polymorphisms and PCa risk, which may provide valuable insights for the risk assessment of PCa.The difference of DNA repair ability (DRC) among individuals is an important factor affecting individual's cancer risk.22.1Published studies on the association between APE1 rs1760944 and rs1130409 gene polymorphisms and PCa risk were restricted to a meta-analysis. Two independent researchers retrieved documents from Science Direct/Elsevier, Embase, CNKI, PubMed, the Cochrane Library and Medline with a search deadline of March 2020 and no language or learning type restrictions. The retrieval terms consisted of MeSH terms and text words. For example, the retrieval terms for the APE1 rs1760944 and rs1130409 gene were “APE1 rs1760944 gene ”, “rs1760944 ”, “APE1 rs1130409 gene”, “rs1130409”, “Apurinic/apyrimidinic (AP) endonuclease”, or “APEX1 gene”; those for PCa were “prostate cancer,” “prostatic neoplasms”, “cancer of prostate”, “cancer of the prostate”, “neoplasms, prostate”, “neoplasms, prostatic”, “prostate neoplasms”, “prostatic cancer”, or “PCa”; and those for the polymorphism were “SNP”, “single-nucleotide polymorphism”, “polymorphism”, “variation”, or “mutation”. All abstracts and relevant documents were retrieved. At the same time, the references in related articles are manually searched, and only full-text documents were included. This study was approved by the Ethics Committee of Chongqing Traditional Chinese Medicine Hospital.2.2Inclusion criteria: If case-control studies tested the relationship between APE1 rs1760944 and rs1130409 gene variants and PCa, the study was included. The case group was PCa patients The criteria for PCa were suspicious findings on a digital rectal examination (DRE) and/or elevated PSA serum levels (>4.0 ng/ml), followed by histopathological confirmation of prostate cancer. And the control groups were normal people. The SNPs of APE1 rs1760944 and rs1130409 were genotyped by polymerase chain reaction-restriction fragment length polymorphism. Effective data were extracted from the article, including eligible and genotype cases and controls, as well as the number of cases and controls of each APE1 rs1760944 and rs1130409 genotype.Exclusion criteria: Case reports, only abstracts, meeting reports, and studies lacking a control population were excluded. Additionally, literature reviews that duplicated previous publications were excluded.2.3Two independent researchers searched the titles and abstracts of the literature, and all relevant studies that met the research criteria were screened. If the title and abstract of the literature could not be used to judge whether the study should be included, the full text was retrieved for analysis. If there were differences in the literature included, it was assessed by consensus or by a third reviewer. In meta-analysis, two reviewers conducted quality assessments based on the main criteria for nonrandomized and observational studies of the Newcastle-Ottawa Quality Assessment scale (NOS).2.42) in the pooled analysis, which represents the percentage of total variation in all studies. If the P value is less than 0.1 or the I2 value is more than 50%, the random effect model is used to analyze and summarize the estimated values. Otherwise, the fixed effect model is applied. The association between APE1 rs1760944 and rs1130409 polymorphisms and the risk of PCa was studied in an allele model (G vs T) , additive model (GG vs TT)  rather than having no minor alleles (TT)) is twice as likely to affect the outcome in a certain direction as is having just one minor allele (TG) rather than no minor alleles (TT)), dominant model (GG and TG vs TT)  vs not having it at all (TT)), recessive model (GG vs TG and GG)  vs having at least one major allele G (TG or TT)), and codominant model (TG vs TT) . The P < .05 suggest significant associations between the APE1 rs1760944 and rs1130409 gene polymorphism and PCa. In addition, publication bias was detected by visual symmetry of funnel plots, with asymmetry suggesting possible publication bias. This was also assessed by the Begg's and Egger's test in the meta-analysis. If the P-value was less than .05, publication bias existed.Data were collected from the literatures by three reviewers and contained demographic statistics  and number of cases and controls for each APE1 rs1760944 and rs1130409 genotype. Differences were settled by consensus. Two researchers conducted quantitative meta-analyses using RevMan and Stata software. The combined odds ratio (OR) and its 95% confidence interval (CI) were calculated. Heterogeneity was assessed using p-values and the I-square statistic . Suggests that APE1 rs1760944 polymorphisms might be a protective factor of PCa, The combined ORs and 95% CIs were as follows: Additive model  . APE1 rs1130409 is suggested to be a risk factor of PCa. The combined odds ratios and 95% confidence intervals were as follows: Additive model    5%, CI 0.9, 0.97)  1.01, 1.5)  Fig. A, 11A, s 1.01, 1.5)  Table B, 11B.4In this study, we focused on seven studies reported to elucidate the relationship between APE1 rs1760944 and rs1130409 gene polymorphisms and PCa risk. Four studies explored the relationship between APE1 rs1760944 gene polymorphism and PCa, the data indicate that APE1rs1760944 gene polymorphism might a protective factor of PCa. For APE1 rs1130409, six studies have studied the relationship between APE1 rs1130409 gene polymorphism and PCa. However, the data indicate that APE1 rs1130409 gene polymorphism might one of the risk factors of PCa.,24 The body's complex DNA repair system has evolved to correct DNA damage caused by the environment. There are many DNA repair pathways, each responsible for repairing specific types of damage. The base excision repair (BER) pathway mainly eliminates DNA damage caused by ionizing radiation, reactive oxygen radicals and methylation agents. Apurinic/apyrimidinic (AP) endonuclease APE1 gene is located at chromosome 14q11.2, which encodes one of the main enzymes in BER pathway and accounts for almost all the cleavage activity of abasic sites observed in cell extract.–32 In addition, APE1 is also a multifunctional protein, which participates in other important cell processes, including response to oxidative stress, regulation of transcription factors, cell cycle control and apoptosis. Therefore, the abnormal expression of APE1 gene may lead to the repair of defects in these lesions and give individuals the susceptibility to cancer.–38 Epidemiological studies have found that APE1 gene polymorphism is related to the risk of lung cancer, breast cancer, colorectal cancer and bladder cancer.,36 18 SNPs were identified in APE1, including two functional SNPs rs1760944 in the promoter region and rs1130409 in the fifth exon. These two functional SNPs have been widely studied. It is reported that APE1 rs1760944 T > G polymorphism is related to the change of promoter activity in vitro, compared with T allele, rs1760944 G allele is related to reducing cancer risk by enhancing transcription activity. APE1 rs1130409 polymorphism T to G, in the biological significance, APE1 codon T to G polymorphism is associated with the delay of lymphocyte mitosis in healthy subjects, suggesting a higher sensitivity to ionizing radiation. And some authors describe the relationship between allele G and prostate cancer. Some studies have analyzed this transition and support this polymorphism as a low penetrance risk factor for cancer development.Human exposure to a variety of environmental carcinogens and endogenous/exogenous reactive oxygen species may lead to DNA damage, which ultimately increases the susceptibility of men to prostate cancer (PCa).Our findings suggest that APE1 rs1760944 gene polymorphism might be a protective factor of prostate cancer. In this meat-analysis the G/G genotype in control group was significantly higher than prostate cancer group. The transformation from codon T to G at this site, may influenced the enhancing the transcriptional activity, protect DNA from damage, in turns reduce gene mutations, resulting in the decrease risk of prostate cancer. For APE1 rs1130409, our findings suggest that APE1 rs1130409 might a risk factor of prostate cancer, in our meta analysis, G/G genotype in prostate cancer group was significantly higher than in control. The possible mechanisms may the APE1 codon T transversion G polymorphism is have higher sensitivity to ionizing radiation, G alleles are more likely to cause gene damage, resulting in the increase risk of prostate cancer. However, studies with larger sample sizes are needed to better illuminate the mechanisms of the APE1 rs1760944 and rs1130409 in the prostate cancer tumorigenesis.There are some limitations in our study, which need to be taken into consideration when interpreting the results of this meta-analysis. First, the sample size of each study was relatively small, and a total of 1769 prostate cancer patients and 2237 normal controls were investigated in the seven studies. Second, several studies on this issue were excluded owing to a lack of control data. Furthermore, because of the limited amount of original research, a subgroup of APE1 rs1760944 and rs1130409 gene polymorphism in different race was not conducted. As such, it is difficult to draw definitive conclusions about the clinical value of APE1 rs1760944 and rs1130409 gene variants inPCa.In summary, the results of this meta-analysis suggest that the current article adds to the evidence of an association between the APE1 gene polymorphisms and prostate cancer progression. These data suggest that APE1 rs1760944 polymorphisms might be a protective factor of prostate cancer, and APE1 rs1130409 is suggested to be a risk factor of prostate cancer. APE1 rs1760944 and rs1130409 polymorphisms may be used in the risk assessment of PCa. However, studies with larger sample sizes are needed to definitively determine the correlation between APE1 rs1760944 and rs1130409 gene polymorphisms and prostate cancer.Conceptualization: Jinnian Liu, Jian Zheng, Yongjian Yin, Shengqiang Qian, Bin Xu, Wei Xiong, Xiangrui Yin.Data curation: Jinnian Liu, Yu Guo, Xia Sheng, Yongjian Yin, Shengqiang Qian, Bin Xu, Xiangrui Yin.Formal analysis: Jinnian Liu, Jian Zheng, Xia Sheng, Yongjian Yin, Wei Xiong, Xiangrui Yin.Funding acquisition: Jinnian Liu, Jian Zheng, Yu Guo, Xia Sheng, Shengqiang Qian, Bin Xu, Wei Xiong, Xiangrui Yin.Investigation: Jinnian Liu, Jian Zheng, Xia Sheng, Xiangrui Yin.Methodology: Jinnian Liu, Yu Guo, Shengqiang Qian, Bin Xu, Wei Xiong, Xiangrui Yin.Project administration: Jinnian Liu, Yu Guo, Xia Sheng, Yongjian Yin, Shengqiang Qian, Bin Xu, Wei Xiong, Xiangrui Yin.Resources: Jinnian Liu, Jian Zheng, Xia Sheng, Bin Xu, Xiangrui Yin.Software: Jian Zheng, Xia Sheng, Xiangrui Yin.Supervision: Jian Zheng, Yu Guo, Yongjian Yin, Shengqiang Qian, Wei Xiong, Xiangrui Yin.Validation: Jian Zheng, Yongjian Yin, Shengqiang Qian, Xiangrui Yin.Visualization: Jinnian Liu, Yu Guo, Xia Sheng, Shengqiang Qian, Wei Xiong, Xiangrui Yin.Writing – original draft: Jinnian Liu, Yu Guo, Xia Sheng, Yongjian Yin, Wei Xiong, Xiangrui Yin.Writing – review & editing: Jinnian Liu, Jian Zheng, Yu Guo, Yongjian Yin, Xiangrui Yin.