It was highly controversial whether fermented dairy foods protect against colorectal cancer (CRC) because of conflicting results from current human epidemiologic studies; we therefore conducted this meta-analysis based on the case–control and cohort studies to estimate the holistic analyses. Finally, a total of seven case–control studies and ten cohort studies comprising a total of >20,000 cases were incorporated in the quantitative synthesis. Specifically, statistical evidence of significantly decreasing CRC risk in case–control studies was found to be associated with cheese intake . In a subgroup analysis, cheese intake was correlated with lower colon cancer  and rectal cancer  risk in case–control studies. Furthermore, we also found that the higher intake of yogurt may lower the risk of rectal cancer  in cohort studies. The consumption of fermented dairy foods may be relevant to decrease CRC risk in this meta-analysis.https://www.crd.york.ac.uk/prospero/display_record.php?ID=CRD42021269798, CRD42021269798. In 2020, colorectal cancer (CRC) was the third most leading cause of cancer death in the Western world. Results of epidemiological studies show that a multitude of risk factors are relevant to colorectal cancer, including a lifestyle, diet, genetics, and obesity, and diet played a pivotal role for the disease . There wIncreasing data had supported a role for the imbalances gut microbiota played on colorectal carcinogenesis . Multiplhttps://www.crd.york.ac.uk/prospero/display_record.php?ID=CRD42021269798 to evaluate the impact of the fermented dairy foods intake on colorectal cancer (CRC).In this study, yogurt and cheese were chosen typically represented on fermented dairy foods in order to ascertain the concrete links in colorectal cancer. We developed this comprehensive meta-analysis on published cohort and case–control studies according to the PROSPERO guidelines Three databases  were searched for all articles in English language since database inception in July 2021. We employed the following terms in the analysis: “fermented food or fermented milk or cultured milk or cheese or yogurt or lactic acid bacteria” and “Colorectal Neoplasm” or “Colorectal Cancer” or “Colonic Cancer” or “Rectum Cancers” and so on. The study design was not restricted during the retrieval of process in order to gain a comprehensive search of literature.If a study meets the following criteria, the results could be incorporated for inclusion in the meta-analysis: (1) a topic of the association about yogurt or cheese consumption and CRC, colon, or rectal cancer risk; (2) the outcome relied on dietary information from questionnaires; (3) odds ratio (OR), hazard ratio (HR), and relative risk (RR) with 95% CIs can be acquired through reading full-text articles; (4) original articles were published in English; and (5) the study of design was a cohort or case–control study. In addition, we may exclude some articles that meet our exclusion criteria. (1) Duplicate articles in different databases; (2) cell or animal experiments; (3) meta-analysis studies; (4) reviews, letters, and commentaries; and (5) articles lacking specific data.First of all, duplicate literature from the databases was removed from this study. Then, two authors (ZL and JW) independently screened the titles and abstracts to exclude some works of data that did not meet our eligibility. In parallel, based on the inclusion and exclusion criteria, the outcome of specific information can be acquired through reading the full texts, such as first author of the works, year of publication, sex, country of recruitment, follow-up period, dairy type, and the number of cases or controls.The Newcastle–Ottawa Scale (NOS) was performed for some cohort or case–control studies. Two reviewers (ZL and JW) determined the quality of the included studies independently. The Newcastle–Ottawa Scale (NOS) of the maximum score was 9, and a high score (≥6) indicated high quality in this study. If there were any discrepancies, disagreements were addressed through discussion.2 statistics. If I2 ≥ 50% from the statistics analysis, the fixed-effects model was performed for calculation; otherwise, the random-effects model was employed. In order to explore the sources of heterogeneity, we also performed a sensitivity analysis by logistic meta-regression analyses. In addition, we examined prespecified stratified analyses for different study characteristics: region, dairy type, sex, and tumor location . The funnel plot and Begg’s rank correlation method were employed to assess publication bias.We calculated the consumption of yogurt or cheese in the highest compared with the lowest categories to computer odds ratios or (case–control studies) rr and hr (cohort studies) corresponding to the 95% confidence interval (95% CI). The study of heterogeneity was assessed with the Cochran’s Q statistic and IThe statistical analyses were performed by STATA 16.0 software .The flow diagram of the steps was presented as a flowchart in The main characteristics of the studies are depicted in Yogurt The outcome of the high consumption compared with low consumption on yogurt is shown in Cheese The outcome of the high consumption compared with low consumption on yogurt is shown in 2 > 50%.). In order to comprehensively analyze the outcome, we formed subgroups on sex, countries, and tumor location.We will consider heterogeneity among studies in overall comparisons and choose the random-effects model (P heterogeneity< 0.001 and IIn order to understand the meta-stability of the associations observed, we omit one study at a time from the outcome in cohort and case–control studies. If the observation did not appreciably change, we confirmed the reliability of the data analysis.The publication bias of selecting literature was evaluated by Begg’s test. Here, we developed this meta-analysis on 17 cohorts and case–control studies with more than 1,310,276 participants and 14,944 cases. By comparing the high consumption with low consumption in two distinct study designs, we attempted to identify the link of fermented food such as yogurt and cheese in association with colorectal cancer. We discovered new associations, some of which have not previously been published. In general, whereas our results from this meta-analysis are consistent with the observation that the consumption of yogurt and cheese remained unclear connection for previous studies, we discovered some novel links in different study designs and subgroups.For example, we observed that different primary tumor sites were associated with yogurt consumption. Results of the case–control study found yogurt intake to be associated with a decreased risk of rectal cancer, but not colon cancer. The reason for this disparity in the outcome was unclear. Furthermore, we found that there is an inverse association between dietary intake of cheese in case–control and the risk of colorectal cancer. There are several reasons that could account for these outcomes. Compared to other fermentation products such as yogurt, having greater viable probiotics in cheese is advantageous. The reason for these differences was interpreted with unique characteristics in higher pH and buffering capacity and lower oxygen and salt levels. In these settings, the long-term survival of probiotics was observed in the center of the cheese. Theoretically, it may also play a protective role in storage and passage through the gastrointestinal tracts . From thvia multiple mechanisms (Epidemiologic studies have shown that fermented dairy foods such as yogurt and cheese, the main sources of probiotics in human diets, have proved to be one source of calcium in Western populations . Therefochanisms Figure 5chanisms –40. HencIn this meta-analysis, there were some inadequacies, although we enrolled a great deal of high-quality studies. No apparent publication bias has not been perceived, yet its impact may remain. Besides this, there was a marked heterogeneity throughout this analysis. The reason could be attributed to the difference in food products, regular and prolonged dietary habits, and the sample size of this study. Finally, the smaller sample sizes of the relevant analysis allows us not to fully account for mortality in colorectal cancer. We hope more experimental and theoretical evidence will be able to verify the outcome.In conclusion, our meta-analysis suggests that fermented dairy food intake may have an impact on the incidence of colorectal cancer. Besides, the economic approach applied to convey health benefits by way of modifying the gut microbiota has been used to ferment dairy foods, which could markedly prevent colorectal cancer in the near future. It may thus be an effective strategy to integrate fermented dairy foods into eating habits for the early prevention of colorectal cancer. In parallel, we wished to see what role this meta-analysis could play in the dietary management of future outbreaks in colorectal neoplasms.The raw data supporting the conclusions of this article will be made available by the authors, without undue reservation.LL , JW, and XS conceived and designed the study. ZL and JH selected the studies and collected the data. RW, PL, and ZD analyzed data. All authors interpreted the results. ZL and JW drafted the paper. All authors revised the draft paper. All authors contributed to the article and approved the submitted version.The authors declare that the research was conducted in the absence of any commercial or financial relationships that could be construed as a potential conflict of interest.All claims expressed in this article are solely those of the authors and do not necessarily represent those of their affiliated organizations, or those of the publisher, the editors and the reviewers. Any product that may be evaluated in this article, or claim that may be made by its manufacturer, is not guaranteed or endorsed by the publisher.


Research on neutrophil-to-lymphocyte ratio (NLR), platelet-to-lymphocyte ratio (PLR), and monocyte-to-lymphocyte ratio (MLR) in depression is still emerging and has increased 3-fold since the first meta-analysis. An updated meta-analysis with sufficient studies can provide more evidence for a potential relationship between NLR, PLR, MLR, and depression.We identified 18 studies from the PubMed, EMBASE, Cochrane library, and Web of Science databases. Meta-analyses were performed to generate pooled standardized mean differences (SMDs) and 95% confidence intervals (CIs) between patients with depression and controls. Sensitivity analysis, subgroup analysis, meta-regression, and publication bias were conducted.p < 0.001 and SMD = 0.24, 95% CI: 0.02–0.46, p < 0.05, respectively). MLR was slightly higher in depressed individuals compared to controls , despite the absence of significance. Sensitivity analysis removing one study responsible for heterogeneity showed a higher and significant effect  of MLR. Three subgroup analyses of NLR, PLR, MLR, and depression revealed obvious differences in the inflammatory ratios between depressed patients and controls in China and the matched age and gender subgroup. Individuals with post-stroke depression (PSD) had higher NLR and MLR values as compared to non-PSD patients . Meta-regression analyses showed that male proportion in the case group influenced the heterogeneity among studies that measured NLR values (p < 0.05).A total of 18 studies including 2,264 depressed patients and 2,415 controls were included. Depressed patients had significantly higher NLR and PLR compared with controls (SMD = 0.33, 95% CI: 0.15–0.52, Higher inflammatory ratios, especially NLR, were significantly associated with an increased risk of depression. In the subgroup of China and matched age and gender, NLR, PLR, and MLR were all elevated in depressed patients vs. controls. Individuals with PSD had higher NLR and MLR values as compared to non-PSD patients. Gender differences may have an effect on NLR values in patients with depression. Major depressive disorder (MDD) is a common and severe mental illness characterized by prominent and persistent low mood, cognitive impairment, and impaired social function . MDD hasThere is increasing evidence to suggest that the immune and inflammatory system plays a role in the pathophysiology of depression –6. IndivNeutrophil-to-lymphocyte ratio (NLR), platelet-to-lymphocyte ratio (PLR), and monocyte-to-lymphocyte ratio (MLR) are inexpensive and easily acquired reproducible markers of systemic inflammation –19, and A previous meta-analysis reported that patients with MDD had higher NLR compared with healthy controls . HoweverWe performed this systematic review and meta-analysis based on the Preferred Reporting Items for Systematic Reviews and Meta-analysis (PRISMA) standard .We conducted a comprehensive literature search in the PubMed, EMBASE, Cochrane library, and Web of Science electronic databases. The following search strategy was used: (“neutrophil-to-lymphocyte ratio” OR “NLR” OR “neutrophil to lymphocyte ratio” OR “neutrophil/lymphocyte ratio” OR “neutrophil-lymphocyte ratio” OR “neutrophil lymphocyte ratio”) AND (“platelet-to-lymphocyte ratio” OR “PLR” OR “platelet to lymphocyte ratio” OR “platelet/lymphocyte ratio” OR “platelet-lymphocyte ratio” OR “platelet lymphocyte ratio”) AND (“monocyte-to-lymphocyte ratio” OR “MLR” OR “monocyte to lymphocyte ratio” OR “monocyte/lymphocyte ratio” OR “monocyte-lymphocyte ratio” OR “monocyte lymphocyte ratio”) AND (“depression” OR “affective” OR “mood” OR “depressive disorder” “major depression” OR “major depressive disorder” OR “MDD”). The search period was from the creation of the databases up until August 2021. We also screened references from selected articles to confirm potentially related studies.Studies were included in this meta-analysis if they satisfied the following criteria: (1) study design: observational studies ; (2) participants: patients who were clearly diagnosed as depressed according to the Diagnostic and Statistical Manual of Mental Disorders (DSM) or International Classification of Diseases (ICD), and controls without mental disorders; (3) comparative data on NLR, MLR, and PLR between cases and controls could be extracted. The exclusion criteria were: (1) duplicate data; (2) reviews, letters, conference abstracts, or meta-analyses; (3) non-human studies and mechanism studies; (4) non-English publications; and (5) non-adult participants. Furthermore, the latest or the most complete study was selected when participants overlapped in several studies. Two researchers (Yanwei Cheng and Yiwen Wang) worked independently to complete the preliminary screening through browsing titles and abstracts, and made final decisions after reviewing the full texts. Disagreements between the researchers were resolved by group discussions.Two investigators (Yanwei Cheng and Xiangyi Wang) extracted original data from the selected studies. The following information was collected: surname of the first author, publication year, country, type of depression, diagnostic criteria, sample size, mean age, percentage of men, and peripheral blood cell ratios. When the data on peripheral blood cell ratio were expressed as being in the median and interquartile range (IQR), we referenced the method of McGrath et al.  to estimThe quality of the selected investigations was independently evaluated by two authors (Yanwei Cheng and Zhuoya Jiang) with the Newcastle-Ottawa scale (NOS). We assessed a study with respect to three aspects: Selection, Comparability, and Exposure or Outcome. A maximum of one star was awarded in each category for Selection and Exposure or Outcome, while a maximum of two stars could be awarded for Comparability. The total score ranged from 0 to 9, and a study could be qualified as high-quality when the total score was not <5. Disagreements were managed by group discussion.p-value < 0.05 was considered statistically significant for the pooled SMDs measured by Z-test. Heterogeneity across studies was measured based on the p-value of the chi-square and I2-statistic. Heterogeneity was considered high if the p-value was < 0.10 or I2 > 50%, and random-effects models were used. A fixed-effects model was applied when the p-value was ≥ 0.10 or I2 ≤ 50%. We conducted a sensitivity analysis in order to investigate the influence of a single study on the overall result, and to search for whether a study is the source of heterogeneity. Subgroup analysis was carried out to seek the potential source of heterogeneity, and to explore whether the estimated effect could vary in the following aspects, country of origin , type of depression , diagnostic criteria (DSM or ICD), matching age and sex simultaneously (Yes vs. No), suicidal behavior including suicide attempt and suicidal ideation (Yes vs. No). Further, we performed meta-regression analyses to identify potential sources of heterogeneity by adjusting mean age and the male proportion. In addition, publication bias was assessed by the funnel plot, Begg's test, and Egger's test.The meta-analysis was carried out using STATA 12.0 statistical software. Standardized mean differences (SMDs) and 95% confidence intervals (CIs) were used to construct forest plots of continuous data and to estimate the association of NLR, PLR, and MLR with depression susceptibility. A A flowchart representing the screening procedure is shown in The eighteen studies –56, 102 p < 0.001) . Sensitivity analysis revealed that the corresponding SMD values did not obviously change after sequential exclusion of each study, both in terms of direction and significance of the results  . The ran results .The results of subgroup analysis based on clinical characteristics  are presented in p = 1.000 and p = 0.165, respectively). These results indicated the absence of publication bias.Meta-regression analyses showed that the differences in mean age and proportion of males in the control group were not the possible sources of heterogeneity. However, we observed a significant difference, namely that the male proportion in the case group influenced heterogeneity. The results of meta-regression are exhibited in p < 0.05), and the forest plot is shown in I2 = 88.0%, p < 0.001), the random-effects model was applied. Relevant sensitivity analysis after sequentially removing several studies displayed the lower and meaningless pooled SMD values . The results suggested no publication bias in reporting the values of PLR in depressed patients and controls.As shown in p > 0.05)  and the random-effects model was used. Sensitivity analysis eliminating one study  . Heterogne study  exhibitene study , which sThere were three subgroup analyses for comparison of MLR values in depressed individuals and controls . The samThe updated meta-analysis including 18 studies suggested that higher peripheral blood inflammatory ratios, especially NLR, were significantly correlated with an increased susceptibility to depression. This result is consistent with that of the previous meta-analysis on MDD performed by Mazza et al. . But ourNeutrophils are front-line troopers of innate immune defense, and engage in a variety of pathological and physiological processes such as injury, inflammation, autoimmunity and so on . Their cFor lymphocytes, they are the main effectors of adaptive immunity which facilitate pathogen-specific immune recognition, immune memory production, and regulation of host immune homeostasis  Figure . Lymphoc+ T cells  in peripheral blood from MDD patients  from B cells .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.All claims expressed in this article are solely those of the authors and do not necessarily represent those of their affiliated organizations, or those of the publisher, the editors and the reviewers. Any product that may be evaluated in this article, or claim that may be made by its manufacturer, is not guaranteed or endorsed by the publisher.


I2 = 0%) with no indication of publication bias. No association of alcohol consumption was found with mortality risk in PCa patients  and PCa mortality risk in healthy subjects . In conclusion, this study suggests that there is some evidence of an association between high alcohol consumption and an increased risk of incidence of fatal prostate cancer in healthy subjects. Given the inconsistencies this result warrants further confirmation.We conducted a systematic review and meta-analysis to investigate the role of alcohol consumption with the prognosis of prostate cancer (PCa). Published reports were gathered on 15 October 2022, from PUBMED/MEDLINE and EMBASE. We found 19 independent eligible studies on the association between consumption of alcoholic beverages and the risk of fatal PCa (n = 5), PCa mortality (n = 5) in healthy subjects, and PCa patients’ survival (n = 7) or surrogates thereof (n = 2). We used random effects meta-analysis to obtain a summary risk estimate (SRE) and 95% confidence intervals (95%CI) for incidence of fatal PCa and PCa mortality. The meta-analysis revealed no association between alcohol consumption and fatal prostate cancer incidence risk in healthy subjects with an indication for publication bias, but omitting the study that mainly increased the between-study heterogeneity, the SRE becomes significant , and the heterogeneity disappeared ( Alcohol is considered as a carcinogen in humans by the Agency for Research on Cancer  and reprProstate cancer (PCa) is the second most common cancer among men worldwide and it ranks fifth in terms of mortality, as it was estimated to have caused over 375,000 deaths in 2020 . ConcernAlcohol consumption has also been investigated in relation to its possible effect on cancer mortality and the prognosis of cancer patients. To date, evidence on the relationship between alcohol consumption and survival or mortality in cancer patients, and particularly in PCa patients, is limited. In an ecological study, Penuelas and colleagues found a significant and positive association between per capita intakes of alcohol and mortality from total, colon, lung, breast, and prostate cancers . In anotDespite the recognized detrimental effects of ethanol for human health and its known effect on morbidity and mortality , alcoholhttps://osf.io/2rvkq/) accessed on 28 December 2022).The protocol of this systematic review is available online in the Open Science Framework (OSF) registry website  criteria illustrated in Concerning the exposure, we considered: (1) intake of alcoholic beverages , either categorized  or quantified in terms of drinking frequency ; (2) alcohol intake , either categorized  or quantified in terms of frequency  or actual amount ; (3) history of alcohol abuse .For estimates included in the meta-analysis, we converted the alcohol consumption categories for comparison in grams per day. We used information reported in the articles, and if missing data occurred, we consider one drink defined as one 12-fluid-ounce beer, one 5-ounce glass of wine, or one 1.5-ounce shot of liquor, all equaling approximately 13 g of alcohol .Mendelian randomization studies considering genetically predicted alcohol intake or variants in genes involved in alcohol metabolism as the exposure of interest were also considered as eligible for inclusion in the review, although they were not entered in meta-analysis models.In terms of outcome, we included papers that reported on any of (1) incidence of fatal PCa  among healthy subjects; (2) PCa mortality ; (3) overall and cause-specific survival of PCa patients, or (4) clinical surrogates thereof, e.g., post-diagnosis changes in PSA level including biochemical failure .p-values, if those were not available), as well as the distribution of potential confounders of the studied association  and variables that were used for adjustment in statistical analyses. For the meta-analysis when more estimates were available, we considered the one regarding highest vs. lowest exposure and sensitivity analyses were conducted when the choice was not straightforward.Upon removing duplicate items, all papers were initially screened out based on the title an abstract, and those that were retained were read in full text in order to decide on their eligibility. The selection of articles was conducted independently by two authors (SC and EP), and any disagreement was resolved by consensus with a third author (MF). The following information was extracted by two researchers (SG and OD) from each paper and entered in a dedicated spreadsheet: first author, year of publication, period of enrolment, country, study design, sample size, risk estimates and their 95% confidence intervals , all the studies included in the review have an observational design: therefore, the study quality and susceptibility to bias was assessed by using the Newcastle-Ottawa quality scale (NOS)  and the The main characteristics of the studies included in the review were reported in tables and qualitatively summarized in narrative form. Meta-analysis was conducted by fitting random effect models, which output a summary risk estimate (SRE) and corresponding 95% CI for the association between high vs. low alcohol consumption. Risk estimates were pooled by means of meta-analysis for each outcome of interest for which we have at least three estimates with a similar exposure, i.e., risk estimates for alcoholism and from mendelian randomization were not pooled with the other estimates.When more estimates from a single study had to be included, a hierarchical model was adopted to take into account that the estimates were not independent.2 index and the Chi-squared test. The 2I index can be interpreted as a measure of the percentage of the total variability of risk estimates across studies that is accounted for by actual heterogeneity instead of mere chance; as a thumb of rule, values of 2I exceeding 50% are considered as an indicator of large between-study heterogeneity. Even when heterogeneity was large, however, subgroup analyses and/or meta-regressions could not be conducted due to the low number of included studies. We carried out sensitivity analyses in order to investigate whether between-study heterogeneity was due to single studies and whether conclusions depended on the choice made among the available estimates.Statistical heterogeneity was evaluated through the IPossible publication bias was evaluated by means of Egger’s linear regression and Begg’s correlation test ,27. Furtp values were two sided and p < 0.05 was considered statistically significant. Meta-analyses were carried out using the Rstudio software (R version 4.1.1 R Core Team (2020). R: A language and environment for statistical computing. R Foundation for Statistical Computing, Vienna, Austria)).All reported The literature search in PUBMED and EMBASE returned 2702 items, which dropped to 2181 after duplicates were removed . A totalThe main results and study characteristics of the selected studies are presented in Five studies reported on the association between alcohol intake and the risk of fatal PCa , of whicp-value 0.007), Begg’s test  and the trim and fill analysis did not detect any potentially missing studies as no asymmetry was revealed from the funnel plot.Although Egger’s regression showed that there was publication bias (2 83%) was entirely attributable to the study by Watters et al. [2I 0%), and there was no indication of publication bias according to both Egger’s regression  and Begg’s test . Finally, the results did not change by replacing the risk estimate for beer consumption with that for liquors in Hsing’s study [p = 0.13).The sensitivity analysis showed that the large heterogeneity . There was no indication that publication bias was at play , and the trim and fill method also confirmed that there were no potentially missing studies as no asymmetry was revealed from the funnel plot. The meta-regression showed that study quality is not an effect modifier (p = 0.97).The meta-analysis comparing the highest vs. lowest category of alcohol consumption confirmed the lack of an association between alcohol consumption and PCa mortality risk in healthy subjects . The heterogeneity across studies was not large (I2 13.8). Using the trim and fill method, potentially missing studies were found, since the funnel plot revealed asymmetry, but did not change the results .No publication bias was reported . Ly et al., reported occurrence of biochemical failure and Burton et al. the growth of PSA concentration . In bothAlcohol consumption categories for comparison in grams per day for estimates included in the meta-analysis are presented in The study quality was generally fair ; the maiIn particular, with regard to the incidence of fatal prostate cancer scenario, the article with the lowest quality score is Huynh Le et al. The latter lacks adequate cohort follow-up, good outcome assessment, exposure assessment and does not make clear whether the outcome of interest was absent at the start of the study. On the other hand, studies on prostate cancer mortality show high quality scores. Among studies reporting surrogates of prostate cancer patient survival: the study by Ly et al. is at high risk of bias due to possible confounding factors not considered in the statistical analysis. While in studies reporting the overall and cancer-specific survival of prostate cancer patients, the works showing the greatest risk of bias are those of Yu et al. and Brunner et al.In this systematic review and meta-analysis overall, we found no significant association between alcohol consumption and incidence of fatal prostate cancer (five studies) or prostate cancer mortality (five studies) in healthy subjects. In studies conducted among PCa patients, our meta-analysis showed that there was no association between alcohol consumption and overall or cancer-specific survival (seven studies). Finally, both studies that examined PCa patients’ alcohol intake in relation to clinical parameters known to correlate with survival, found no association with the corresponding surrogate of survival. Indeed, the results of individual studies included in our review are mixed and inconsistent.Regarding the relationship between alcohol consumption and fatal PCa incidence, only one study  includedWith respect to PCa mortality among healthy subjects, two studies included in our review reported a significant association with alcohol consumption, but with opposite results: Breslow et al. found an increased risk of PCa mortality among higher-frequency drinkers (1–2 or ≥3 days per week vs. less than once weekly), while the study by Dickerman et al. showed that abstainers had a greater risk of death from PCa compared to light drinkers (0.01–3 drinks/week) ,36. DickMoreover, with regard to alcohol consumption among PCa patients, two studies included in our review ,39 foundOur study has some limitations that are appropriate to mention. First, the validity of our results is curbed by the low number of available studies for each of the outcomes under scrutiny, by their methodological limitations, and above all by their heterogeneity. The main source of heterogeneity lies in the way the exposure of interest was defined and measured, since the included studies quantified the risk of death from all causes and from specific causes in healthy subjects or PCa patients in relation to different exposures such as lifetime (or never) alcohol use, alcohol intake after PCa diagnosis, alcoholism/alcohol dependence syndrome, or the presence of variants in alcohol metabolizing genes. In addition, as we have already mentioned, there may be misclassification errors in the reference categories. For example, sicker men who had drunk in the past and stopped drinking at baseline, could be misclassified attenuating the association.In this meta-analysis we could not study dose–response relationships of alcohol and prostate cancer outcomes due to missing information. Subsequent research should investigate this aspect.The lack of association reported by studies can be due to the possible effect of uncontrolled confounding factors such as diabetes history. The use of antidiabetic medications such as metformin, can exert a protective effect, reducing risk of prostate cancer as reported in the literature ,50. The In conclusion, this systematic review and meta-analysis indicates that there is some evidence that alcohol consumption may play a role in fatal prostate cancer risk. Alcohol consumption has been linked to an increased likelihood of developing over 200 health conditions, not only cancers and cardiovascular diseases, but also liver diseases, road injuries and violence, suicides, tuberculosis and HIV/AIDS . For the


Evidence regarding associations of circulating saturated fatty acids (SFAs) with chronic diseases is mixed. The objective of this study was to determine the associations between total or individual SFA biomarkers and the risk of cardiometabolic diseases.Four electronic databases were searched from inception to March 2022. Three investigators independently assessed for inclusion and extracted data. Random-effects or fixed-effects models was used to estimate the pooled relative risks (RRs) and corresponding 95% confidence intervals (CIs) for the association of total or individual SFA biomarkers, including even-chain SFAs , odd-chain SFAs  and very-long-chain SFAs , with risk of incident type 2 diabetes (T2D), cardiovascular disease , CHD and stroke.A total of 49 prospective studies reported in 45 articles were included. Higher concentration of circulating total SFAs was associated with an increasing risk of cardiometabolic diseases, the risk increased significantly by 50% for CVD (95%CI:1.31–1.71), 63% for CHD (95%CI:1.38–1.94), 38% for stroke (95%CI:1.05–1.82), respectively. Similarly, levels of even-chain SFAs were positively associated with higher risk of chronic diseases, with RRs ranging from 1.15 to 1.43. In contrast, the risk of cardiometabolic diseases was reduced with increasing odd-chain SFA levels, with RRs ranging from 0.62 to 0.91. A higher level of VLCSFAs corresponded to 19% reduction in CVD. Further dose-response analysis indicated that each 50% increment in percentage of total SFAs in circulating was associated with an 8% higher risk of T2D  and trends toward higher risk of CVD . Inverse linear relationships were observed between 17:0 biomarker and T2D or CVD risk.Our findings support the current recommendations of reducing intake of saturated fat as part of healthy dietary patterns. Further studies are needed to confirm our findings on these SFAs in relation to cardiometabolic outcomes and to elucidate underlying mechanisms.https://www.crd.york.ac.uk/prospero/display_record.php?ID=CRD42022329182], identifier [CRD42022329182].[ In vitro and animal studies have demonstrated that SFAs can accelerate the atherosclerotic process, induce cellular inflammation and influence the insulin sensitivity (Limiting intake of saturated fatty acids (SFAs) has been considered as a key component to reduce the risk of chronic disorders for decades . The 202sitivity , 5. Resusitivity . TherefoAssociations of dietary SFAs with the incidence of major cardiometabolic diseases have been reported by some large prospective cohort studies, and findings of these studies were inconsistent –9. InterTo fill this knowledge gap, a dose-response meta-analysis of data from prospective studies was conducted to evaluate the associations between the circulating even-chain SFAs, odd-chain SFAs, VLCSFAs as well as the total SFAs and the incident of T2D, CVD, CHD, and stroke.This study was registered in PROSPERO (CRD42022329182) and reported in accordance with the Preferred Reporting Items for Systematic Review and Meta-Analysis (PRISMA) guidelines.A comprehensive literature search of Web of Science, PubMed, EMBASE, and Cochrane Library were conducted to identify relevant published articles from inception to March 2022. The following terms were used: “fatty acids” or “saturated fatty acid” or “saturated fatty acids” or “lauric acid” or “myristic acid” or “pentadecanoic acid” or “palmitic acid” or “margaric acid” or “stearic acid” or “arachidic acid” or “heneicosylic acid” or “behenic acid” or “tricosylic acid” or “lignoceric acid,” AND “type 2 diabetes” or “diabetes mellitus” or “impaired glucose” or “impaired fasting insulin” or “cardiovascular disease” or “coronary heart disease” or “heart disease” or “ischemic heart disease” or “coronary artery disease” or “myocardial infarction” or “stroke” or “ischemic stroke” or “haemorrhagic stroke,” AND “blood” or “marker” or “biomarker” or “serum” or “plasma” or “whole blood” or “adipose tissue” or “circulating” or “erythrocytes” or “red blood cell” or “cholesterol esters,” AND “prospective” or “nested case-control” or “cohort” or “case-cohort” or “follow-up” or “longitudinal.” The search was restricted to human studies without any language restriction. Additionally, reference lists of retrieved articles, review articles, and meta-analyses were manually scanned to identify any other relevant studies. Corresponding authors of eligible articles were contacted for any further relevant work, published or unpublished, to reduce risk of publication bias.Studies were included in the meta-analysis if they satisfied the following criteria: (1) the study design was prospective  with a follow-up more than 1 year; (2) the exposures of interest were total or individual SFA biomarkers  measured in any type of tissue [circulating blood  or adipose tissue]; (3) the endpoints of interest included incident T2D, CVD, CHD, and stroke; (4) the adjusted relative risks (RRs), odds ratios, or hazard ratios with corresponding 95% confidence intervals (CIs) or standard errors were presented or could be calculated. Whenever reports referred to identical outcomes from the same population, only those with the highest number of cases or the longest follow-up times were retained to avoid data duplication. Potentially eligible studies were assessed independently by three investigators , with discrepancies resolved by discussion until consensus was reached.Data extraction were conducted independently by 3 investigators . A standardized data collection form was applied to extract the following baseline characteristics from each study: first author name, publication year, study location, study name, study design, follow-up duration, participant characteristics , SFA subtype, biological sample type , assessment method , outcomes of interest (type and number of cases), covariates adjusted for in the multivariable model, and the risk estimate with 95%CIs for all categories of each biomarkers. When studies reported multiple results based on different numbers of covariates included in statistical adjustments, the results that adjusted for the most number of variables were extracted.Study quality was performed by the same authors according to the validated Newcastle-Ottawa Scale (NOS) , which aI2 statistic , prospective case-cohort study (PCC), or nested case-control study (NCC)], geographical location , assessment method , biomarker type , number of cases (< 300 or ≥ 300), as well as study quality (moderate or high). Dose-response analyses were conducted based on the method proposed by Greenland and Longnecker , GLC (n = 18), NMR (n = 2), and GC-tandem mass spectrometry (n = 1). 48 studies measured SFA biomarkers in circulating blood  and one in adipose tissue. Most studies were adjusted for age (n = 47) and smoking (n = 46), and alcohol drinking (n = 40), physical activity (n = 39) and body mass index (n = 38) were also controlled for in many studies. The results of the study quality assessment showed that 44 studies were graded as high quality, others were graded as moderate quality  higher risk of T2D. For every 50% increase in percentage of odd-chain SFAs concentration in total SFAs, the risk of T2D was reduced by 8% for 15:0 , 16% for 17:0 , respectively.Twenty-two studies were assessed to estimate risk between the concentrations of SFA biomarkers and T2D, including 166,745 participants and 21,252 T2D events , 27–44. P < 0.001; Pheterogeneity = 0.56; P = 0.002; Pheterogeneity < 0.001; P = 0.03; Pheterogeneity < 0.001; P < 0.001; Pheterogeneity = 0.15; P = 0.001; Pheterogeneity = 0.45; P = 0.09; Plinearity = 0.96) for each 50% increment in percentage of total SFAs in circulating. For every 50% increment in percentage of 17:0 and 24:0 concentrations of total SFAs, the RR of CVD decreased by 18% for 17:0 , 10% for 24:0 , respectively.Association between circulating SFAs and risk of CVD were investigated in 27 studies, with 53,845 participants and 14,613 CVD events , 45–66. P < 0.001; Pheterogeneity = 0.45; P = 0.05; Pheterogeneity = 0.003; P = 0.02; Pheterogeneity = 0.21; P = 0.03; Pheterogeneity = 0.22; P < 0.001; Pheterogeneity = 0.23; A total of 14 studies with 7,314 cases in 26,772 participants were included to investigate the relationship between circulating SFAs and CHD risk , 49–59. P = 0.02; Pheterogeneity = 0.76; P < 0.001; Pheterogeneity = 0.11; P = 0.004; Pheterogeneity = 0.10; The association between circulating SFAs with stroke risk were investigated in nine studies, comprising 16,589 participants and 3,176 stroke events , 60–66. P > 0.05 for both tests).Sensitivity analysis showed that no single study substantially changed the statistical significance or direction of the combined RR for any of the outcomes. Visual inspection of the funnel plot did not reveal substantial asymmetry –67. SimiIn the present meta-analysis, higher concentration of circulating total SFAs was associated with an increasing risk of CVD, CHD, and stroke, rather than T2D. Compared with the adverse effect of even-chain SFAs, odd-chain SFAs and VLCSFAs showed a significant inverse association with these cardiometabolic diseases. Our study suggested that individual circulating SFAs may not be equally associated with the risk of chronic diseases, which raises the possibility of generating public health recommendations and nutritional guidelines of SFAs in these cardiometabolic diseases prevention.Previous studies have found that higher dietary intake of SFAs was associated with an increasing risk of chronic health condition . Data frin vitro experiments, Pal et al. found that even-chain SFAs could induce proinflammatory cytokine expression in adipocytes through the FetA and TLR4 pathway, resulting in insulin resistance  stress –73. Systsistance . Animal sistance . Along wsistance , 78. Thesistance . Clinicasistance , 81. Measistance , which msistance  and initogenesis , ultimatSD increment of the sum of 15:0 and 17:0 was associated with a 10% decrement of this inflammatory marker (Compared with the deleterious effects of even-chain SFAs, the present results suggested that a higher level of odd-chain SFAs was associated with a lower risk of chronic conditions. Data from Northern Sweden Health and Disease Study suggested that each percent increase of the proportion of this fatty acid species was significantly associated with a risk reduction of a first myocardial infarction . The Väsy marker . Additioy marker . Moreovey marker , 88–93. y marker , 95. DesFor VLCSFAs, the present study of an inverse association between circulating VLCSFAs and CVD risk was in line with the results from previous epidemiological studies , 96. WhiSome limitations of this study should be mentioned. Firstly, individual blood SFAs were quantitatively measured once and changes of these SFAs over time might lead to potential random measurement error caused by within-person variation. However, among the Nurses’ Health Study and Nurses’ Health Study II participants, reasonable validity and reproducibility of concentration of SFAs such as 15:0, 16:0, and 24:0 have been demonstrated with intraclass correlation coefficients, indicating a single measurement of fatty acid biomarker levels can reliably represent long-term levels over time . SecondlOur meta-analysis of existing prospective studies indicated that a higher concentration of total SFAs and even-chain SFAs was associated with an increasing risk of cardiometabolic diseases, which supported the current recommendations of reducing intake of saturated fat as part of healthy dietary patterns. In addition, our finding of protective effects of odd-chain SFAs has potentially important clinical implications for preventing cardiometabolic diseases. Further studies are apparently needed to confirm our findings on these SFAs in relation to cardiometabolic outcomes and to elucidate underlying mechanisms.The original contributions presented in this study are included in the article/LM, JL, and MM generated the idea for the study, formulated an analytical plan, and supervised the study. ZL, HL, and HJ searched for relevant studies, conducted quality assessment, and extracted the data. ZL and HJ performed data analysis. ZL and HL wrote the manuscript and all other authors revised the manuscript. All authors approved the final manuscript for submission and acquired, analyzed, or interpreted the data.


The prevalence of colorectal cancer (CRC) and inflammatory bowel disease (IBD) is increasing worldwide. Understanding what factors foster these diseases’ development can help reduce their burden. Viral infections have been suggested to be involved in the genesis of these diseases, but the data is contradictory. The present study analyzed 11,413 articles on the topic and identified 196 that could provide information on the relationship between a viral infection and these gastrointestinal diseases. The Epstein–Barr virus  was strongly associated with IBD, and cytomegalovirus (HHV-5) with both CRC and IBD.Gastrointestinal diseases (GDs) include colorectal cancer (CRC), gastric cancer (GC), and inflammatory bowel disease (IBD). CRC and GC are typically diagnosed at later stages of development, reducing patients’ chances of survival. IBD is characterized by chronic intestinal inflammation and is a significant risk factor for the development of CRC. Chronic bacterial infections have been shown to promote some GDs, but the role of viruses in the etiology of these diseases is less clear. The present meta-analysis retrieved literature on the viral prevalence in GD patients, measuring the GD risk in odd ratios. By quantifying the study heterogeneity, the literature bias was fundamentally included in the analysis. The analysis also included 11 metagenomic studies. Our meta-analysis retrieved 11,413 studies, with 196 suitable for analysis. HHV-4 (Epstein–Barr virus) was identified as a significant risk factor for the development of IBD, and HHV-5  as a risk factor for both CRC and IBD. Polyomaviruses and the Hepatitis B virus were also, less strongly, involved in the risk of CRC and IBD. No relations withstanding the literature bias were identified for GC. The study discusses these findings, as well as the role of other viruses in the etiology of CRC and IBD. About 15% of all human cancers have been shown to have a microbial etiology . In someCancer of the stomach  and colorectal cancer (CRC) share some similarities, including a high number of deaths caused worldwide, detection (at least in the western world) at rather late stages of the tumor development, and risk factors such as hereditary mutations, obesity, and sedentariness . CRC is Every person has a four percent life-time risk of developing sporadic CRC , but somHelicobacter pylori is known as a major risk factor, albeit the precise oncogenic mechanism is still poorly understood, and only a subset of H. pylori carriers develop GC [Fusobacterium nucleatum might put further strains on the aberrant activation of NF-κB, further enhancing the gastrointestinal carcinoma risk [Recently, specific infections have been suggested to affect the risk of sporadic CRC . For insvelop GC . It is bvelop GC . An infeoma risk  and a HHoma risk . Still, oma risk .The idea of a relation between infection and CRC has been discussed in the last few decades . SeveralCaudovirales, has been consistently reported [In relation with the establishment of the current paradigm linking oncogenic bacteria and CRC , bacterireported .Both CRC and GC develop rather slowly and symptoms are noticed only when the tumors have grown to a considerable size . Thus, aIs the presence of viruses associated with an increased risk of developing CRC/GC/IBD? We answered this question by surveying the literature in the public domain and stratifying the eligible articles into groups that were used to determine the odds ratios (ORs) able to indicate the strength of such an association.Given the inconclusive data present in the public domain, we sought to provide an updated meta-analysis regarding the viral associations with CRC, GC, and IBD, with particular attention put on the phages. The research question posed herein was the following: Adenoviridae: Adenovirus (ADV).Anelloviridae: Transfusion Transmitted Virus .Astroviridae: Astrovirus (AstV).Bunyaviridae: members of the Orthobunyavirus genus (BV).Caliciviridae: Norovirus (NV), Sapovirus (SaV).Herpesviridae: Human Herpesvirus 1 (HHV-1 or Herpes Simplex virus), Human Herpesvirus 3 , Human Herpesvirus 4 , Human Herpesvirus 5 , Human Herpesvirus 6 (HHV-6), Human Herpesvirus 8 , Inoue–Melnick virus (IMV).Hepadnaviridae: Human Hepatitis B virus (HBV).Flaviviridae: Human Hepatitis C virus (HCV).Matonaviridae: Rubella virus (RuV).Retroviridae: Human Immunodeficiency virus (HIV), Human T-cell lymphotropic virus type 1 (HTLV-I).Papillomaviridae: Human Papilloma virus (HPV).Parvoviridae: Parvovirus B19 (B19), Bocaparvovirus (HuBV).Polyomaviridae: BK virus (BKV), John Cunningham virus (JCV), Merkel cell polyomavirus (MCPV), Polyovirus 5 (PyV6), Simian virus 40 (SV40).Paramyxoviridae: Measles virus (MeV), Mumps virus (MuV).Reoviridae: Reovirus (RV), Rotavirus (RoV).Syphoviridae: phages of the order Caudovirales.Myoviridae: phages of the order Caudovirales.Inoviridae: member of the Inovirus genus.The present study was conducted following the Preferred Reporting Items for Systematic Reviews and Meta-Analyses (PRISMA) . The comHerpesviridae).Herpesviruses .Respiratory viruses (Papillomaviridae).Papillomaviruses .Epithelial viruses .Intestinal viruses (Polyomaviridae).Polyomaviruses (Hepadnaviridae).Hepadnaviruses (Flaviviridae).Flaviruses (Retroviridae).Retroviruses .Phages  without infection, number of cases with infection, number of healthy patients (controls) without infection, number of controls with infection, tissue type, analytical method and virus species, were tabulated for further analysis. The articles were also subdivided into virome and non-virome studies. The studies were further stratified, based on the source of the data: the tissues were obtained from the colon-rectum, stomach , stools, sera, and medical  records.I² index [I² index were stratified, as suggested [The assessment of the literature bias was measured, taking the study heterogeneity (variation of the degree of association between the cause and effect) into account, and quantified with the Higgins I² index . The valuggested : low (<3uggested . The ERTuggested . HoweverThe odds ratio (OR) statistics were calculated for the non-virome studies, using the random effects model implemented by Mantel and Haenszel . The triI² with their 95% CI and associated p-value. The classes that did not contain data were not reported in the table to increase the readability. The most relevant data were also displayed in forest plots and funnel plots to increase the understanding of the underlying relations between the studies [The data were summarized in the tables reporting the OR and  studies ,46.Representativeness: if the selected article included more than 50 participants, a cut-off accepted to provide the statistically solid results, a positive value was assigned.Control group: if the selected article included a control group, a positive value was assigned.Documentation: if the selected article was not a letter, a positive value was assigned.Outcome: if the selected article was not a virome study, a positive value was assigned.Co-infections: if the selected article did not report samples with co-infections, a positive value was assigned.The risk of bias for the observational studies was assessed with the Newcastle–Ottawa scale (NOS)  by prepa5 points: High3–4 points: Moderate1–2 points: LowThe sum of the assigned points was used to stratify the selected studies as follows:For the virome studies, it was possible to do a qualitative description only, which was due to the absence of the clear number of cases and the controls infected with a given virus. Thus, the viral prevalence is provided for these cohorts as a whole.R version 4.2.1 using the packages metafor, meta, and metasens, which provided both the statistical and graphical outputs.The meta-analysis was implement in We inquired six scientific literature databases: PubMed, Cochrane Library, Web of Science, Cinahl, ClinicalTrial.gov, and the International Clinical Trials Registry Platform (ICTRP) on 3 March 2021 and identified 11,413 articles, with 1009 duplicated entries, matching our search algorithm a. FollowSince there were 38 articles of the 196 studies retrieved (19.4%) which reported results for more than one virus, there were collectively 255 data points, 211 of them (82.7%) performed after the year 2000 b. There The data points could be stratified, according to the following groups:CRC .GC .IBD .Hybridization . Fluorescent immuno-assay (IFA), in situ hybridization (ISH) or immuno-histochemistry (IHC).PCR . Polymerase chain reaction (PCR), both end-point or quantitative.Archival . Medical records obtained from the retrospective studies, and without direct detection of the virus within the specimens.Serology . Enzyme-linked immunoassay (ELISA) and plaque assay.Virome . Whole-genome sequencing (WGS) by massively parallel sequencing (MPS).Colon and rectum tissue .Gastrointestinal (stomach) tissue .Stools .Sera .Medical records .Respiratory viruses .Intestinal viruses .Epithelial viruses .Herpesviruses .Retroviruses .Polyomaviruses .Papillomaviruses .Hepadnaviruses .Flaviviruses .Phage group .The majority of the 196 studies displayed a moderate NOS score , followed by a high NOS score , and only a fraction of the studies retrieved herein had a low score . Therefore, it could be assumed that the quality of the dataset was good .p-value = 0.0187), Polyomaviruses , Hepadnaviruses , and epithelial viruses . An infection with Papillomaviruses was just above significance . Based on the sera, none of the studies indicated a significant increase of risk of a gastrointestinal disease. Conversely, based on the archival records, the random effects model identified a significant association between infection with Flaviviruses  and confirmed the association between an infection with Hepadnaviruses and CRC . However, the results obtained from the archival records was based on individual studies and pointed to a decreased risk of cancer upon infection. None of the indices related to the quantification of study heterogeneity  could be computed, confirming the low reliability of the results, based on archival data.The risk of developing either CRC, GC or IBD due to a viral infection based on the random effects model, is reported on p-value = 0.1803), suggesting a negligible publication bias.Similarly, the results regarding the increased risk of CRC upon infection with HBV, based on the colorectal tissues, was based on a single study and even in this case no measure of heterogeneity could be computed. The association between the increased rick of CRC and infection with epithelial viruses was slightly better because, based on two studies that consistently reported a higher cancer risk upon infection, only the CQT could be computed (I² index was considerable (≥70.3%), indicating an issue of publication bias and corroborating the use of the random effects model for computing the meta-analysis because it is more reliable in the presence of such a bias. The ERT was slightly below significance, strengthening the issue of publication bias. The observation of the funnel plots for Herpesviruses, Papillomaviruses, and Polyomaviruses, showed that the reports related to an infection with HPV were seriously affected by an asymmetric distribution , whereas the CQT became non-significant. Given the small sample set, the former metric should be considered more reliable. However, none of the viruses was associated with a significant risk of CRC; thus, the increased risk of CRC, based upon an infection with either Herpesviruses and Polyomaviruses, could not be confirmed by the trim-and-fill method.Given the influence of the publication bias in defining the association between viral infection and CRC, the trim and fill method was applied . The ERTHerpesviridae family . HHV-5 was instead just above significance . HHV-1 (Herpes Simplex 1) and HHV-8 (Kaposi sarcoma herpes virus), with only one study each, did not provide valuable data to the analysis.Further stratification of the CRC risk, based on a Herpesviruses infection, was based on the viral genera of the e family . Based oI² = 83.3%) but moderate for HHV-5 (I² = 60.1%), albeit the CQT was significant in both cases. The ERT could not be calculated for either sub-group. The application of the trim-and-fill method resulted in the addition of two studies with both HHV-4 and HHV-5. For the former, the association produced the values: OR = 1.24, 95% CI: 0.19–8.21, p-value = 0.8268, I² = 84.3% (95% CI: 69.3–91.9%), CQT < 0.0001; for the latter the values were: OR = 1.70, 95% CI: 0.55–5.25, p-value = 0.3573, I² = 64.5% (95% CI: 24.3–83.4%), CQT = 0.0062. The ERT could not be computed for either sub-group. Therefore, even in this occurrence, the fill-and-trim method did not substantiate the results of the random effects model. A visual inspection of the data  was due to JCV . The other viruses involved in the infection model were BKV , SV40 , and MCPV . Further analysis was focused on JCV because it was the most prevalent in the study set. The random effects model reported a significant association, based on 15 studies, between an infection with JCV and the CRC risk: OR = 2.64, 95% CI: 1.27–5.49, p-value = 0.0096, I² = 82.9% (95% CI: 73.0–89.1%), ERT = 0.0524, CQT < 0.0001. However, the trim-and-fill method, which added five studies to the set, did not confirm the association: OR = 1.19, 95% CI: 0.46–3.09, p-value = 0.7180, I² = 87.3% (95% CI: 81.8–91.2%), ERT = 0.7247, CQT < 0.0001.The vast majority of the studies related to the CRC risk and an infection with members of the The epithelial group associated with a high risk of CRC, was due to an infection with TTV (n = 2) and B19 (n = 1). Due to the paucity of data, it was not possible to further stratify the analysis. The other sources of information  did not provide additional value to the analysis of the CRC risk.p-value = 0.3567), resulting in a non-significant association and the absence of heterogeneity measures. Only two studies provided information regarding the GC risk upon infection with Polyomaviruses, but in this case with a significant association: OR = 5.22, 95% CI: 1.66–16.47, p-value = 0.0048, ERT = 0). It was not possible to perform the trim-and-fill method.The number of studies regarding GC was very limited. Based on the gastric tissues, only Herpesviruses and Polyomaviruses had enough coverage to provide metrics for the meta-analysis . Howeverp-value = 0.2774; Hepadnaviruses: OR = 1.36, 95% CI: 0.72–2.57, p-value = 0.3375.Serum samples provided information for Herpesviruses and HBV, but none of the associations was significant , CQT = 0.3980. The trim-and-fill method did not add any study to the set, thus it confirmed the results of the random effects model. The moderate I² score and the non-significant CQT, highlighted that in this case, there was a negligible publication bias and, therefore, no requirement for an adjustment of the results.Based on the colorectal tissues, the only viruses providing information to compute in the meta-analysis, were the Herpesviruses, which, however, showed an association just above significance: OR = 2.71, 95% CI: 0.91–8.06, p-value = 0.0236. However, the heterogeneity was substantial , suggesting that there was a strong publication bias. The trim-and-fill method added three articles to the set and did not confirm the association: OR = 0.69, CI: 0.19–2.45, p-value = 0.5630, I² = 94.2% (95% CI: 90.8–96.4%), CQT < 0.0001. Remarkably, the adjustment of the data showed that the IBD risk could be as high as 2.45 upon infection, suggesting that the possible beneficial role of an infection with intestinal viruses  was most probably a statistical artifact. With only one study supporting the association between an infection with respiratory viruses (Adenovirus) and the IBD risk, the association was even weaker than for intestinal viruses: OR = 0.29, CI: 0.07–1.16, p-value = 0.5630. Neither the measurement of the study heterogeneity nor the trim-and-fill method could be computed.Based on stool specimens, intestinal and respiratory viruses showed a significant association with the IBD risk. There were five studies substantiating the meta-analysis, however the trend reported was that of a reduction of IBD upon infection with intestinal viruses: OR = 0.27, 95% CI: 0.09–0.08, p-value = 0.0241, I² = 72.3% (95% CI: 40.2–87.2%), CQT = 0.0014. The trim-and-fill method added one study and confirmed the association between an infection with Herpesviruses and the high risk of IBD: OR = 2.67, 95% CI: 1.10–6.52, p-value = 0.0307, I² = 69.1% (95% CI: 35.6–85.2%), CQT = 0.0019. Based on gastric tissues, there was a significant involvement of Herpesviruses in increasing the IBD risk. The association was supported by seven studies and provided the following measures: OR = 2.90, 95% CI: 1.15–7.53, p-value = 0.0291. Neither the measurement of the study heterogeneity nor the trim-and-fill method could be computed. As in the case of the association between an infection with respiratory viruses and the IBD risk based on stools, the reduced risk upon infection might simply be a statistical artifact. Similarly, the association between an infection with Polyomaviruses and the IBD risk, was based on a single study: OR = 9.47, 95% CI: 3.46–25.92, p-value < 0.0001. In this case, there was an elevated risk but the paucity of the study set made the association untrustful. Neither the measurement of the study heterogeneity nor the trim-and-fill method could be computed.Based on the sera, there was a significant association between a high IBD risk and an infection with either respiratory viruses and Polyomaviruses. The former association was, however, based on a single study: OR = 0.38, 95% CI: 0.16–0.91, p-value = 0.2492, I² = 90.1% (95% CI: 77.6–95.6%), CQT < 0.0001. However, the trim-and-fill method added two studies and reported a significant association: OR = 0.55, 95% CI: 0.32–0.94, p-value = 0.0275, I² = 94.5% (95% CI: 90.6–96.8%), CQT < 0.0001 , CQT = 0.3740. The trim-and-fill method did not add any study, confirming the negligible publication bias associated with these measures , CQT = 0.1761. The trim-and-fill method added two studies but the association was reduced to just above significance: OR = 2.26, 95% CI: 0.88–5.82, p-value = 0.0920, I² = 56.4% (95% CI: 0–83.8%), CQT = 0.0571  than in the healthy controls (n = 12) ; however, such an increase did not correspond to a rise in the phage richness [Caudovirales family), together with a lower bacterial richness in the IBD patients (n = 52) over the healthy controls (n = 21), was instead reported by Norman and co-workers  [Lactococcus, Lactobacillus, Clostridium, Enterococcus and Streptococcus, as specifically associated with this disease.The majority of the virome studies were based upon stool samples: 8 out of 11 (72.7%). The microbiome analysis showed a higher viral richness in the controls (n = 28) than the CD patients (n = 31): there were 4399 viral species in the former group and 2161 viral species in the latter ,49. Syne lesions . Other srichness . An incr < 0.05) . In partp-value = 0.013) as well as an inverse association between the bacterial and viral richness [Orthobunyavirus as a taxon specifically associated with CRC specimens, followed by Inovirus and Tunalikevirus. Using a similar approach, Hannigan and collaborators [Siphoviridae and Myoviridae  as more associated with CRC (n = 60) than the healthy condition (n = 30) .The comparison of the viral communities between the CRC cases (n = 74) and the healthy controls (n = 92) in the stools, showed an increased phage richness in the former .In one study based on colon biopsies , the phages B40-8 and B124-14, both infecting  lesions . In one revalent . HealthyThe work by Zapatka and collaborators was based on the re-analysis of the stored WGS-data, derived from 5354 pairs of tumors and normal tissues . The tumGastrointestinal diseases represent a primary public health concern. In particular, CRC and GC are highly prevalent cancer types with a high mortality and delayed diagnosis. IBD is a significant cause of morbidity that has been pointed out as a risk factor for the development of CRC. The prevalence of CRC and GC is still increasing. A genetic predisposition, dietary habits, and sedentary lifestyle, amongst other causes, are associated with developing these gastrointestinal diseases. Bacterial infections have been recognized as additional risk factors as well. However, the role of viral infections in the development of CRC, GC, and IBD is less well understood. Our present meta-analysis suggests a role for HHV-4 (EBV) as a risk factor for IBD and also, less significantly, for CRC, and of HHV-5 (CMV) as a less strong risk factor for CRC and IBD.The present study sought to assess the role of viruses as risk factors for these main gastrointestinal diseases. This aim was fulfilled by applying a meta-analysis to the scientific publications related to the prevalence of viruses in people with CRC, GC, or IBD, against control subjects. We included all viruses associated with human diseases, including bacteriophages, to avoid the risk of missing possible links.The most striking finding provided by the present study is the amount of literature bias on this subject. Among the causes of study heterogeneity, there might be, e.g., geographical variations or the use of disparate investigative methods . In the The data presented by our meta-analysis confirm the role of the particular viral infections as risk factors for CRC and IBD. The primary finding of the current meta-analysis was that, using both the random effect model and the trim-and-fill model, the OR of an infection over the absence of an infection was significantly greater than the unity, indicating that the Herpesvirus HHV-4 (EBV) was a significant risk factor in the development of IBD. The present study’s second finding was less accurate than the primary result, since it was based on the random effect model only. Still, it suggests HHV-5 as a risk factor for both CRC and IBD. The meta-analysis supported the tertiary findings that were not only based on a single model, but also hampered by a small sample set and, consequently, a high publication bias. Such results suggested that: (a) Polyomaviruses could be associated with a higher risk of both CRC and IBD; (b) HBV could be associated with a higher risk of CRC; (c) Papillomaviruses could not be ruled out as CRC risk factors since the association was just slightly above significance; and (d) respiratory, intestinal, and epithelial viruses had a very limited prognostic value on the risk of these. In contrast, there was not sufficient evidence to provide an association between a viral infection and the GC risk, due to the absence of a high enough number of strong studies.Herpesviridae in colorectal (HHV-4), gastric (HHV-4 and HHV-5) and liver  tissues [HHV-4 (EBV) is typically associated with lymphomas, rather than gastrointestinal diseases, whereas HHV-5 is believed to cause chronic inflammation and immune dysfunction in immune-compromised people ,57. Howe tissues . These f tissues . Remarka tissues , suggest tissues . In this tissues ,63.3K, and MAPK) [It is well known that HHV-4 can disrupt the cell cycle at several points, fostering the genesis of cancer and even metastasis ,64. For nd MAPK)  and redund MAPK) , which ind MAPK) . LMP2A ind MAPK) . Additiond MAPK)  and the nd MAPK) . Based ond MAPK) . The oncnd MAPK) .HHV-4 (EBV) has been shown to be more prevalent in IBD patients ,74. For IBD is a CRC precursor . Since ITherapy with immune suppressants, such as infliximab, is also involved in the reactivation of Hepadnaviruses . The maiAs in the case of Herpesviruses and Hepadnaviruses, Polyomaviruses can also be reactivated by immunosuppressive therapy ,94,95. IAbout a decade ago, HHV-4 was described as a causative agent of the Epstein–Barr virus-positive mucocutaneous ulcer (EBVMCU), a condition affecting the pharyngeal and gastrointestinal mucosa, with ulcers refractory to treatment and displaying histological features typical of HHV-4 associated lymphomas ,107. EBVTaken together, these findings suggest that intestinal damage can be caused by viruses that induce chronic inflammation without directly residing in the tissues of the gastrointestinal tract. HHV-4 might reside in the immune cells, HHV-5 in several tissues, and HBV in the liver, still able to affect the gastrointestinal tract by causing a generalized chronic inflammation and aberrant immune response. This framework might explain why these viruses are not consistently found in CRC and IBD cases. Another explanation for the lack of a clear relationship between a viral infection and CRC, might be that viruses can cause ‘hit-and-run’ mutagenesis ,114. RemThe present review included the analysis of the MPS-derived studies as well. These, being qualitative in essence, could not be used for the quantification of the risk of gastrointestinal diseases upon infection. Although MPS-based studies provide a broader overview of the presence of microbes in samples, they are, at the moment, not suitable for quantitative analysis because they are not designed to calculate ORs. The statistical methods applied in these studies are designed to indicate trends in the presence of microbes in the samples, but not to reveal a measured risk associated with a disease. Nevertheless, the MPS-based studies, including our own, consistently reported a higher viral richness (the number of species) in the inflamed or tumor tissues . A weakeConversely, MPS-derived studies including our recent one, reported a decreased bacterial richness as well ,51,117. Orthomyxoviridae family, did not appear in the results. While a comprehensive review of the relationship between all types of viruses and intestinal diseases may never be complete, the lack of reference between an infection with viruses, such as orthomyxoviruses and CRC, GC, or IBD, suggests that the link is tenuous. Third, the lack of longitudinal data made it impossible to determine whether the infection was a cause or result of intestinal diseases. The presented study only reported a statistical association between an infection and disease; more detailed studies will be required to determine the causality comprehensively.The present study had some limitations. First, the present meta-analysis considered each viral group as a whole regardless of the detection method employed. However, such heterogeneity might have introduced a bias in the results. Ideally, the prevalence obtained with one technique should be confirmed by another method, but this is rarely the case. Herein, the subdivision of the results by detection method, would have reduced the number of studies in each class, hampering their statistical evaluation. The present study aimed to identify the possible trends in the association between a viral infection and intestinal disease that should be further investigated at the experimental level. Second, it was also impossible to include every type of virus. Despite using the general term “virus” as a keyword during the literature search, viruses known to cause gastroenteritis during the acute phase of infection, such as the The present study investigated the prevalence of viruses by a meta-analysis of the actual status of relevant scientific publications in CRC, GC, and IBD. The meta-analysis highlights a significant publication bias in the literature related to this topic. Taking this into account, the present study suggests that HHV-4 (EBV) is a significant risk factor in the development of IBD, and points to HHV-5 as a risk factor for both CRC and IBD. Furthermore, the meta-analysis suggests that, to a less significant extent, Polyomaviruses might play a role in the risk of both CRC and IBD, and that HBV might be a risk factor for CRC.


Colorectal cancer risk and survival have previously been associated with telomere length in peripheral blood leukocytes and tumor tissues. We quantitatively assessed these associations through a systematic review and meta-analysis. Following PRISMA guidelines, we identified relevant studies through database searches, and performed meta-analyses using random effects models. We found no association between telomere length in circulating leukocytes and the risk of developing colorectal cancer, however, shorter leukocyte telomeres were associated with a worse survival in patients with colorectal cancer. Therefore, telomere length may serve as a potential biomarker especially for colorectal cancer prognosis. Larger prospective cohort studies are needed to further confirm this potential association.1/Quartile2–4 overall HR (95% CI) = 1.41 (0.26–7.59) and 0.82 (0.69–0.98), respectively]. (4) Conclusions: Although there was no association with colorectal cancer risk, a poorer survival was observed among those with shorter leukocyte telomere length. Future larger studies evaluating a potentially non-linear relationship between telomeres and colorectal cancer are needed.(1) Background: Colorectal cancer risk and survival have previously been associated with telomere length in peripheral blood leukocytes and tumor tissue. A systematic review and meta-analysis of the literature was conducted. The PubMed, Embase, and Web of Science databases were searched through March 2022. (2) Methods: Relevant studies were identified through database searching following PRISMA guidelines. Risk estimates were extracted from identified studies; meta-analyses were conducted using random effects models. (3) Results: Fourteen studies were identified  through systematic review. While no association was observed between circulating leukocyte telomere length and the risk of colorectal cancer [overall OR (95% CI) = 1.01 (0.82–1.24)], a worse survival for those with shorter telomeres in leukocytes and longer telomeres in tumor tissues was observed [Quartile Telomeres are the nucleotide repetitive structures (TTAGGG) at the end of eukaryotic chromosomes that are sheltered by a protein complex ,2. ShortSeveral studies have suggested that shorter telomeres measured in circulating blood leukocytes are a risk factor for cancer development including colorectal cancer ,17. Two Previous studies that looked at the association between telomere length and colorectal cancer survival have either utilized peripheral blood leukocytes for circulating telomere length measurement or have measured telomere length in preserved tumor tissues recovered during surgery. Two previously published meta-analyses suggested poorer colorectal cancer outcomes among patients with longer telomere length in tumor tissues and shorter telomere length in peripheral blood leukocytes ,23. HoweThe overall aim of this systematic review and meta-analysis was to evaluate telomere length measured in peripheral blood leukocytes as a potential predictive biomarker for colorectal cancer risk, as well as to assess the potential for telomere length measured in circulating leukocytes or tumor tissues to serve as a prognostic biomarker among patients with colorectal cancer.https://doi.org/10.17605/OSF.IO/VRHJP (accessed on 20 January 2023)). The meta-analysis followed the ‘Preferred Reporting Items for Systematic Reviews and Meta-Analyses’ (PRISMA) guidelines  1.01 [0.82–1.24]) .p = 0.03), I2 statistic = 54% and τ2 statistic = 0.03, and a wide prediction interval of 0.59 to 1.73 (Similar results were observed for the comparisons of quartiles 2 and 3 with Q1 . We obse to 1.73 . When dr to 1.73 .We did not observe any significant heterogeneity by study design (retrospective vs. prospective studies) .The association between telomere length and colorectal cancer survival is summarized in We observed contrasting associations with colorectal cancer survival for leukocyte vs. tissue telomere length in our meta-analyses; shorter telomeres in circulating leukocytes were associated with a worse survival (HR [95% CI] 1.41 [0.26–7.59]), although this was not statistically significant, while shorter telomeres in tissues were associated with a better survival after colorectal cancer diagnosis (HR [95% CI] 0.82 [0.69–0.98]) . We wereOur results did not suggest an association between leukocyte telomere length and colorectal cancer risk after meta-analysis of eight studies evaluating colorectal cancer risk. In this meta-analysis of five studies on telomere length with colorectal cancer survival, we observed a worse overall survival with shorter telomeres in circulating leukocytes and longer telomeres in tumor tissue. No evidence of publication bias was observed, though we observed some heterogeneity in previously published studies.To the current meta-analysis, we were able to add two recent studies on telomere length and colorectal cancer risk ,30 that Consistent with two previously published meta-analyses from 2016 and 2017, we reported that longer telomeres in tumor tissues and shorter telomeres in leukocytes were associated with a worse overall survival after colorectal cancer ,23. The Comprehensive studies on telomere-related gene expression in colorectal cancer have reported extreme shortening of telomere length in early-stage colorectal cancer that is compensated by the overexpression of telomere maintenance mechanisms ,46. OverDifferent DNA extraction methods  were reported in the studies included in this meta-analysis. This may have contributed to some of the heterogeneity observed between studies. It has been previously reported that telomere lengths on DNA extracted through column-based systems are shorter compared with telomere lengths measured on DNA extracted using other systems . SimilarOur study is limited by the small number of available studies evaluating this research question, the differences in DNA extraction and telomere length measurement methods in these studies, as well as a lack of consistency in reporting results for telomere length measurement. This observation is consistent with a recent report by Lindrose et al. 2021, that concluded that there is a lack of rigorous reporting of telomere measurement procedures in the published literature . We recoOur meta-analysis includes recently published studies that have not been included in previous systematic reviews or meta-analyses ,30,39. AWe observed no association between telomere length in circulating leukocytes and the risk of developing colorectal cancer. We observed a possible association of shorter telomeres in circulating leukocytes, although it was not statistically significant, and of longer telomeres in tumor tissues with survival after colorectal cancer diagnosis. Thus, telomere length in circulating leukocytes and tumor tissues may have a potential for being prognostic biomarkers for colorectal cancer survival and may aid clinicians to identify patients that have a higher risk for adverse clinical outcomes. Heterogeneity among the included studies was observed, likely due to the limited number of included studies, differences in DNA extraction and telomere measurement methods, as well as a lack of the standard reporting of telomere parametrization. Future studies evaluating the relationship of telomeres with colorectal cancer risk and survival should follow the guidelines for telomere length measurement and reporting, to evaluate the potential of telomeres as predictive and prognostic biomarkers for colorectal cancer.


P < 0.001; I2 = 98%). We conducted subgroup analyses of Chinese  and non-Chinese studies . The sensitivity analysis by risk of bias continued to demonstrate an association, and the heterogeneity remained unchanged . Higher AIP values were associated with higher odds of developing CAD.Various studies suggest that the atherogenic index of plasma (AIP) is associated with the risk of coronary artery disease (CAD) in different clinical scenarios. This review aimed to synthesize evidence of the association between AIP values and CAD. A literature search was carried out on four databases, namely, PubMed, Scopus, Web of Science, and Ovid-Medline. A handsearch was performed on preprint repositories (MedRxiv and Research Square). The effect measurements were expressed as odds ratios (OR) with their corresponding 95% confidence intervals (CI). For the quantitative synthesis, we employed a random-effects model. We analyzed 14 articles  from seven different countries. The quantitative analysis revealed that an increase in one unit of AIP was associated with higher odds of developing CAD (OR 2.11; 95% CI 1.65–2.69;  In Cardiac risk stratification is necessary to improve preventive and therapeutic interventions . BiomarkHypertriglyceridemia and low HDL levels are independent markers of CHD ,16. TherThus, several studies have suggested that AIP acts as a marker associated with cardiovascular diseases in patients with various pathologies ,20,21,2222.1We conducted this systematic review in accordance with the guidelines of the Preferred Reporting Items for Systematic Reviews and Meta-Analysis (PRISMA) statement . The sum2.2We conducted a literature search on articles that evaluated the association between AIP and CAD on October 15, 2021. Four databases were searched: PubMed, Scopus, Web of Science, and Ovid-Medline. A handsearch was also performed on preprint repositories (MedRxiv and Research Square) (see Supplementary Appendix 1).2.3The articles collected from the aforementioned databases were exported to the Rayyan data management software , and dupOnce this phase was completed, each of the articles was read in full-text, and we determined whether they met the previously established selection criteria. A study was excluded if it failed to comply with the selection criteria. If an article did not contain the required information, an attempt was made to contact the corresponding author of the said article. A secondary bibliographic search was conducted on the articles that reached the full-text review phase.In case of disagreements on the inclusion or exclusion of an article, a third reviewer was consulted. Finally, information of the selected articles was extracted and collated in a standardized Excel 2016 document. The extracted data included author name, country, year of publication, study sample, study design, population characteristics , exposure (AIP mean or median of the whole sample and according to sample stratification), and crude/adjusted association measures . Outcome definitions are presented in Table S2.2.4P-values > 0.1 were considered indicative of no publication bias.Four authors  evaluated the included studies independently using the Newcastle–Ottawa Scale . The stu2.5I2-statistic, where values >60% were considered to indicate severe heterogeneity. A Cochran’s Q test was also conducted, where a P-value < 0.05 was considered statistically significant. A random-effects model was used for the meta-analysis as we anticipated heterogeneity among studies. In the subgroup analysis, we conducted a comparison between associations according to country (Chinese vs non-Chinese studies), study design, CAD definition, and gender (women vs mixed). In the sensitivity analysis, we excluded studies with a high risk of bias. We also analyzed all studies after transforming OR to RR. In addition, we performed a meta-regression, reported as bubble plots, in order to search for heterogeneity sources, due to differences in the clinical and methodological characteristics of the individual studies. Assessed variables were study design, study location, CAD definition, and gender.The quantitative analysis was conducted using the RevMan 5.4 statistical program . The OR and their corresponding 95% confidence intervals (CI) were the only effect measures used; thus, we transformed standardized mean differences to ln[OR] using the Chinn method. RR was converted into OR ,26. Vari33.1The search strategy yielded 975 records. After removing 610 duplicates, 365 articles finally remained. After evaluating the titles and abstracts, we excluded 322 more studies. Then, a full-text review was performed on the remaining 43 articles, which led to the exclusion of 29 more articles. In the end, 14 articles were included in the qualitative analysis and meta-analysis ,38,39,403.2The characteristics of each study are presented in detail in 3.3Table S3 presents the results of the methodological quality assessment. In summary, 10 and 4 studies were classified as having a low and moderate risk of bias, respectively.3.4P < 0.001) (I2 = 98%). Statistically significant associations with high heterogeneity were observed in all subgroup analyses by study location , whereas in the other, all OR values were transformed to RR ; however, the association with high heterogeneity remained in both cases   with sevlocation , study dlocation , definitlocation , and genlocation . Sensitilocation  were incth cases .3.5P = 0.974) , study design (P = 0.959) , CAD definition (P = 0.614) , or gender (P = 0.958) .No significantly associated coefficient was observed in study location (3.6P = 0.956), and the funnel plots did not show asymmetry .There was no indication of small-study effects , whose expression in AIP reflects the complex interactions of lipoprotein metabolism and is useful in predicting plasma atherogenicity . sdLDL hFor the aforementioned reasons, our finding of an association between high AIP values and increased CAD risk is not surprising. In addition to the aforementioned factors, factors associated with specific CAD conditions could also explain our results. In this context, some studies have found an association between AIP and arterial stiffness in normotensive patients , which iContrarily, in a multinational study in seven countries from 2003 to 2015, this association was independent of traditional risk factors . SimilarFinally, some studies demonstrated an association between AIP and the evolution of patients with acute coronary syndromes and myocardial infarction. For example, a Turkish study found that AIP was independently associated with the absence of reflow in patients with ST-elevated myocardial infarction following the primary percutaneous coronary intervention . SurprisThe meta-analysis conducted by Wu et al.  found thDespite the evidence suggesting an association between AIP and CAD, this is a systematic review and meta-analysis that systematize the available evidence. We also conducted sensitivity analyses considering the biases of the studies and the place of origin of the studies, which provided robustness to our results. The findings obtained allow us to suggest the use of AIP as a potential low-cost marker of CAD that will allow health personnel to prioritize management strategies and could even be used as part of predictive scores, replacing markers such as triglycerides. Similarly, the usefulness of this marker is independent of the age of the patients ,62, althOur systematic review has some limitations that must be acknowledged. First, there is evidence of high statistical heterogeneity due to the clinical and methodological differences between the analyzed studies. Second, most studies have not controlled the AIP values for confounding variables that impact the proposed outcome. Sociodemographic lifestyles and comorbidity variables should be considered in future studies to avoid confounding bias. Despite this, the direction of the effect is consistent in most of the reported studies. Third, most studies were conducted in Asia, with few or no studies found in other continents. Therefore, it is important to analyze the AIP value in CAD risk in these populations to establish its usefulness. Fourth, no definitive cutoff points have been reported to set the sensitivity and specificity of the AIP in CAD; thus, it is necessary to address it in future research in various contexts and populations. Finally, our findings indicate that high AIP values are associated with a greater possibility of suffering from CAD; however, with the obtained evidence, we consider that AIP is a risk marker and not a causal risk factor because the causes of lipid alterations are mediated by genetic and environmental factors. The AIP in altered ranges would be the expression of this multicausality and, therefore, a biomarker of the pathophysiological process leading to the appearance of CAD.5This systematic review and meta-analysis demonstrated that patients with higher AIP values have higher odds of developing CAD. However, additional primary studies are warranted to define an optimal cutoff point of AIP to predict CAD.