Recently, several studies explored the association between glaucoma and the risk of stroke, but these results were inconsistent. Therefore, we conducted a meta-analysis to examine this possible association.We conducted a systematic literature search of PubMed, Embase, and Web of Science from inception until February 28, 2022. Random-effects meta-analysis was conducted by generic inverse variance method. Sensitivity and subgroup analyses were performed. The review protocol has been registered with PROSPERO (CRD42022312797).Seven studies  have been published from 2004 to 2017 and included in the meta-analysis. These studies included four retrospective cohort studies, two cross-sectional studies, and one case–control study. Meta-analysis of these data has shown that glaucoma was associated with an increased risk of stroke . Most of the subgroup analyses demonstrated similar results. These findings were stable in sensitivity analyses.We found that glaucoma was associated with an increased risk of stroke. The result suggests that patients with glaucoma need to be assessed the risk of stroke to reduce the incidence of stroke. To better explore the nature of any association, prospective studies that consider the stroke subtypes, sample size, district, and other confounding factors are needed. Glaucoma represents a chronic degenerative optic neuropathy characterized by the dysfunction and loss of retinal ganglion cells (RGCs), progressive degeneration of retinal nerve fiber layer, and cupping of the optic disc. Glaucoma can eventually lead to visual field defects –3.Nowadays, glaucoma is the leading cause of global irreversible blindness, some studies reported that approximately 21 million people afflicted by glaucoma in China in 2020, the prevalence among the middle-aged and elderly in Europe was about 2.93%, and due to the aging population, the number of people with glaucoma will increase to 111.8 million worldwide in 2040 . NeverthStroke is the main cause of death and disability , and itsIn recent years, the relationship between glaucoma and stroke has attracted considerable attention. However, the evidence from observational studies –23 showsThis systematic review and meta-analysis has been conducted in accordance with the Preferred Reporting Items for Systematic Reviews and Meta-Analysis (PRISMA) statement , and thiWe conducted a systematic literature search of PubMed, Embase, and Web of Science. The retrieval time was from database inception to February 2022 to identify observational studies on the association between glaucoma and the risk of stroke, and the reference lists of relevant articles were examined to supplement the search. We used Medical Subject Heading terms in combination with free terms searching, without language restrictions. The detailed search strategies were as follows:  AND . The full search strategy for PubMed is described in online The following inclusion criteria were applied: (1) Observational studies  investigating the association between glaucoma and the risk of stroke, (2) Studies providing unadjusted or adjusted effect estimates, the odds ratio (OR), risk ratio (RR), and hazard ratio (HR) with corresponding 95% confidence interval (CI), if these data were not provided, then they will be calculated from raw data wherever possible, (3) The included sample size has a clear time period. Excluded criteria were as follows: (1) Abstract, systematic reviews, editorials commentaries, protocols, opinion papers, letters, and reports, (2) No comparison group, and (3) Studies without sufficient data.Title and abstract screening were done independently by two reviewers for potential eligibility and assessed full-text articles for final eligibility. Discrepancies about selection were resolved through consultation with a third reviewer.The following variables were abstracted from each study: the first author, publication year, study country, study design, type of glaucoma, type of stroke, study subjects, sample size, mean age or age group, percentage of male, follow-up time or study period, OR, RR, HR (adjusted and unadjusted) with their 95% CI, and adjusted confounding variables. Two reviewers independently compared the data selected and resolved the disagreement through consultation.Two reviewers independently used the Newcastle–Ottawa Scale (NOS)  to assesQ-test was performed to evaluate the heterogeneity between studies, P < 0.10 for the Q-test was considered statistically significant, and an I-squared statistical test was conducted to assess the degree of heterogeneity, I2 = 0% indicated no heterogeneity, 25–50% for low, 50–75% for moderate, and more than 75% for high heterogeneity , and the odds ratios (ORs) and 95% confidence intervals (CIs) were calculated to evaluate the association between glaucoma and stroke. A random-effect, generic inverse variance method of DerSimonian and Laird was used to estimate the pooled OR and 95% CI. As the outcome of interest was relatively uncommon, we considered RR/HR equivalent to OR . When boogeneity . StatistAccording to our initial search strategies, 3,243 potentially relevant articles were identified from three electronic databases, of which 449 were excluded as they were duplicates. A total of 2,769 records were excluded after scanning these titles and abstracts. After removing 18 records based on the full-text reading  was observed in the analysis, and sensitivity analyses indicated that the study from Lee et al. , two cross-sectional studies , and one case–control study . In the subgroup analysis stratified by type of glaucoma, we found a significant positive association between open-angle glaucoma and stroke , normal-tension glaucoma and stroke , and neovascular glaucoma and stroke . In the subgroup analysis stratified by type of stroke, we found a significant positive association between glaucoma and unspecific stroke  and ischemic stroke ; however, no significance was observed for study with hemorrhagic stroke . As we all know, gender, age, and region are also risk factors for cerebrovascular disease. We continued to conduct the subgroup analysis of patients stratified by these three variables. In the subgroup analysis stratified by adjustment for confounders, we found a significant positive association between glaucoma and stroke with adjusted OR  and crude OR , respectively. In subgroup analysis stratified by mean age, we found a significant positive association between glaucoma and risk of stroke under 65 years , no significance was observed for the study with more than or equal to 65 years old . The results of the subgroup analyses based on gender ratio of study participants, study region, and study quality were consistent with the overall pooled results.To explain the high heterogeneity, several subgroup analyses were performed. Details of all subgroup analyses are provided see –8. On su studies , 19 .Exploring an association between glaucoma and stroke could be meaningful, for both are common medical disorders, this gives rise to concern from a public health point of view. The purpose of this research was to synthesize the published observational studies on the association between glaucoma and stroke risk through a systematic review and meta-analysis. To the best of our knowledge, the present study is the first meta-analysis on this topic to date. In this meta-analysis of seven studies involving 362,267 participants, we demonstrated that glaucoma was associated with an increased risk of stroke .Six of the seven articles reported their results after adjusting for confounding factors. Four retrospective cohort studies , 21, 23,The meta-analysis included observational studies that can only provide details about correlations but not the causality. However, the association between OAG and stroke risk may also be explained by shared risk factors. The mechanism of glaucomatous optic neuropathy is unknown, but ischemic damage plays an important role in glaucomatous optic nerve damage . ImpaireConsidering the significant heterogeneity of this study, we plan to use meta-regression and subgroup analysis to explore the source of heterogeneity. Because of the limited number of related studies, we just performed subgroup analyses depending on several methodologic and clinical features. Four retrospective studies , 21, 23,I2 index decreasing from 96 to 68%, revealing it as the source of heterogeneity. This difference may be caused by the following reasons: the average age (54 years old) of this article was little younger than that in other articles (62 years old), in addition to that, this article studied the relationship between normal-tension glaucoma (NTG) and stroke but did not provide a clear type of stroke. Meanwhile, other potential sources of heterogeneity may exist that we failed to identify and test due to the overall small number of studies.Sensitivity analysis and most of the results of the subgroup analysis suggest that these meta-analysis results were not affected by the included study or study characteristics. Thus, this suggested that the results of the meta-analysis were relatively stable and statistically robust. However, this result needs to be treated with caution. Sensitivity analysis revealed no change in the results. But after removing the study by Lee et al., the degree of heterogeneity decreased from high to medium, with the This study has some strengths and limitations that are worth considering. First, owing to the high prevalence and incidence of both glaucoma and stroke globally, the topic of our study is meaningful and clinically relevant. To our knowledge, this is the first meta-analysis to date to explore the association between glaucoma and the risk of stroke with a large sample size , we believe that this systematic review and meta-analysis provides the comprehensive assessment of the association between glaucoma and risk of stroke. Second, we made a broad, strict search strategy and conducted independent screening, full-text review, and extraction by two reviewers to ensure the accuracy and integrity of data. In addition to that, in order not to omit studies that meet the inclusion criteria, “Rey” literature in Embase and Web of Science databases was also searched, and we try our hardest to exclude the low-quality studies by using stringent inclusion criteria. Third, we performed subgroup analysis and rigorous sensitivity analysis to evaluate the association between glaucoma and the risk of stroke. Together these results suggest that our conclusion is reliable and convincing.The results of this meta-analysis are new, but some potential limitations need to be pointed out. First, all studies that we included are observational studies that the confounding factors might be present in the original studies, and this may reduce the credibility of our results. Several cerebrovascular disease risk factors, such as gender, age, hypertension, diabetes, and atrial fibrillation, have been adjusted in most articles. However, potential confounding variables, such as smoking, drinking, body mass index, cerebrovascular disease family history, and other cerebrovascular risk factors, were not adjusted. Furthermore, prospective studies with appropriate controls that adjust for potential confounders are urgently warranted to further examine the association between glaucoma and the risk of stroke. Second, the results may not generalize to other districts given that the geographic distribution of the included results is limited, with most studies conducted in the Asian region. Third, data for glaucoma, stroke, and other comorbidities were collected from a secondary claims database, inaccurate or incomplete information about these records may cause data inaccuracy, in addition to that, due to lack of sufficient data on the severity of glaucoma, the association between the severity of glaucoma and the risk of stroke is still unclear.In summary, we found that glaucoma was associated with an increased risk of stroke. The result suggests that patients with glaucoma need to be assessed the risk of stroke to reduce the incidence of stroke. To better explore the nature of any association, prospective studies that consider the stroke subtypes, sample size, district, and other confounding factors are needed.The original contributions presented in the study are included in the article/MW and NC performed the literature search and drafted the manuscript. MW and B-cS screened the literature. NC, C-BG, and SZ extracted data and assessed the quality. M-JH, Z-BH, and X-yW analyzed and interpreted the data. B-GZ and MW designed the systematic review. MW, B-GZ, and NC were responsible for the research design, data analysis, and manuscript revision. All the authors gave their approval for the submission of the final manuscript.


Current studies on the relationship between XRCC3 rs861539 polymorphism and ovarian cancer risk have been inconsistent. Therefore, we performed a meta-analysis to explore their association.  Six electronic databases  were searched for relevant studies published before December 2021. Meta-analysis, subgroup analysis, sensitivity analysis, and publication bias analysis were performed using Stata software 16.0. Trial sequential analysis (TSA) was performed using TSA 0.9.5.10 Beta software. P = 0.003; GG vs. GA+AA: OR = 0.89, 95%CI = 0.82-0.96, P = 0.004). In an ethnically stratified subgroup analysis, XRCC3 rs861539 was shown to reduce the risk of ovarian cancer in Caucasian in the heterozygote model and the dominant model . In the control source and detection method stratified subgroup analysis, hospital-based studies and PCR-RFLP-based studies were found to increase ovarian cancer risk .  A total of 12 studies were included in 9 literatures, comprising 4,634 cases of ovarian cancer and 7,381 controls. After Bonferroni correction, the meta-analysis showed an association between XRCC3 rs861539 polymorphism and ovarian cancer risk in the heterozygote model and the dominant model (GA vs. GG: OR = 0.88, 95%CI = 0.81-0.96,  This meta-analysis showed a significant association between XRCC3 rs861539 polymorphism and ovarian cancer risk, especially in Caucasians. Large-scale multicenter case-control studies in more different regions will be needed in the future. Ovarian cancer is the most lethal of the gynecologic malignancies, with 314,000 new cases and 207,000 deaths worldwide in 2020, and its morbidity and mortality are increasing year by year [DNA damage caused by endogenous  and exogenous  factors can give rise to genomic instability and somatic mutations , 6. The Although many studies have explored the association between XRCC3 rs861539 polymorphism and ovarian cancer risk –26, the To identify eligible case-control studies, we conducted a comprehensive search across six databases  for studies published up to December 2021. We searched the databases using combinations of the following keywords: “XRCC3 or X-ray repair cross-complementing 3 or Thr241Met or rs861539 or 18067C/T” and “ovarian cancer” and “polymorphism or polymorphisms.” Additional articles were searched by checking the references of relevant studies.The studies included in this meta-analysis need to meet the following criteria: (1) focusing on the association of XRCC3 polymorphism rs861539 with ovarian cancer risk, (2) case-control studies, and (3) providing sufficient information on genotype frequency in the case and control groups. The main reasons for exclusion were as follows: (1) reviews, cell or animal studies, and meta-analyses; (2) information on genotype frequency is unclear or absent; and (3) studies contained duplicate data.Two reviewers independently screened the literature according to the selection criteria, extracted essential data from the included studies, and then cross-checked to reach a consensus. The essential data we extracted included the first author's name, year of publication, country, ethnicity, total number of cases and controls, genotype frequency of the case group and the control group, and value of Hardy-Weinberg equilibrium (HWE). If an article had a case-control series of studies, we treated each study as an individual study in this meta-analysis.The quality of the included studies was assessed using the Newcastle-Ottawa scale , which iP < 0.01 (0.05/5) was considered statistically significant. Chi-square-based Q statistic test and I2 value were used to test heterogeneity between studies. When P < 0.1 or I2 > 50%, significant interstudy heterogeneity was indicated, and random effect model was used for analysis; otherwise, fixed effect model was used instead [P value. P < 0.05 indicates that the meta-analysis may have publication bias. The analysis for the present study was completed by Stata software 16.0 .We assessed the association between XRCC3 rs861539 polymorphism and ovarian cancer risk using odds ratios (ORs) and 95% confidence intervals (CIs) in five genetic models, including homozygote model (GG vs. AA), heterozygote model (GA vs. GG), dominant model (GA+AA vs. GG), recessive model (GG+GA vs. AA), and additive model (G vs. A) , 30. Bon instead . Subgrou instead . Sensitiα = 0.05, two sided), 80% power of study (β = 0.20), and a case-control event proportion calculated from meta-analysis [Meta-analysis may lead to false-positive and false-negative results due to random error and lack of statistical accuracy . We condanalysis . TSA wasAccording to the search strategy, we found 137 articles in the database . And 1 article was identified through searching the references of the eligible. Of these, 129 articles were further excluded due to duplicates (58), no reporting of the genotype (17), unrelated to ovarian cancer (19), reviews and meta-analyses (21), nonclinical studies (4), non-case-control studies (3), conference abstracts (5), and duplicate data (2). Ultimately, 9 articles with 12 studies were included for meta-analysis –26. The P > 0.05).The 12 studies included 4,634 cases of ovarian cancer and 7,381 controls, with 10 studies conducted in Caucasians , 24–26, P = 0.003; GG vs. GA+AA: OR = 0.89, 95%CI = 0.82-0.96, P = 0.004), while no significant association was found in the other three genetic models .The meta-analysis results of XRCC3 rs861539 polymorphism and ovarian cancer risk are shown in P = 0.004) and dominant model . In the subgroup analysis based on control source, hospital-based was significantly increased in the homozygote model . In subgroup analyses based on detection method, PCR-RFLP was associated with a significant increase in ovarian cancer risk in the homozygote model .We implemented subgroup analyses of five genetic models based on ethnicity, control source, and detection method . In the P = 0.732 and 0.087, homozygote model P = 0.193 and 0.028, heterozygote model P = 0.732 and 0.626, recessive model P = 0.064 and 0.012, dominant model: P = 0.945 and 0.325. Egger's test P < 0.05 in both homozygote and recessive models, indicating publication bias existed in the two models. Sensitivity analysis results illustrated that our results are reliable , indicating that the meta-analysis may have obtained false-positive conclusions. Therefore, more case-control trials should be included in the future to confirm the conclusions.In this meta-analysis, two genetic models produced statistically significant in the estimation of correlation between XRCC3 rs861539 polymorphism and ovarian cancer risk. Therefore, we conducted TSA to analyze the heterozygote model (GA vs. GG) and the dominant model (GG vs. GA+AA). As shown in Figures In the past few years, the Cancer Genome Project and the Cancer Genome Atlas have intersected with the field of DNA repair, revealing specific characteristics of DNA damage and DNA repair errors in various cancers , which hA number of studies have reported an association between XRCC3 rs861539 polymorphism and ovarian cancer risk –26. MichFor the overall data, the current meta-analysis showed an association between XRCC3 rs861539 polymorphism and ovarian cancer risk in the heterozygote model and the dominant model. In the subgroup analysis based on ethnicity, we found that XRCC3 rs861539 was associated with ovarian cancer risk in Caucasians. In addition, after stratifying patients separately according to the control source and detection method, hospital-based studies and PCR-RFLP-based studies were found to be more associated with ovarian cancer risk. In these subgroup analyses, we suspect that ethnicity, control source, and detection method may be the sources of interstudy heterogeneity.Compared with previous meta-analyses, this meta-analysis included the most comprehensive research with enough samples for in-depth analysis, but it still had some limitations. Firstly, some studies lacked information on subjects' age, age at menopause, and family history, making it impossible to subgroup these factors, which reduces the comprehensiveness of the results. Secondly, many of the relevant studies have been conducted among Caucasians, with fewer studies involving Asians. Thirdly, the interactions of gene-gene and gene-environment may have influenced our results , 45. FouIn conclusion, the present meta-analysis suggests that the XRCC3 rs861539 polymorphism may be associated with ovarian cancer risk, especially in Caucasians. Considering the limitations of this meta-analysis, large-scale multicenter case-control studies in more different regions will be needed in the future.


Interleukin 10 (IL-10) is associated with the progression of leishmaniasis because it inhibits the leishmanicidal action of macrophages and the production of mediators such as IFN-γ and nitric oxide. Studies have shown that specific polymorphisms are associated with the regulatory role of IL-10 and the development of more relevant clinical forms of leishamaniasis. We performed a systematic review and meta-analysis to determine whether single nucleotide polymorphisms (SNPs) of IL-10 influence the progression of leishmaniasis. The selected articles were read in full and only those consistent with the eligibility criteria were included in our study. Seven studies were eligible according to the inclusion criteria and were included in the present systematic review, but only five were subjected to statistical analysis. The pooled odds ratios showed no significant association between the rs1800871 SNP and the progression of leishmaniasis in all genotype models, including the dominant, recessive, homozygote, heterozygote, and allelic models. Regarding the association between rs1800896 SNP and the progression of leishmaniasis, the pooled odds ratios showed no association under all genotype models. Hence, IL-10 SNPs did not show significant association and were not considered a risk factor for the progression of leishmaniasis. Leishmania and transmitted by the bite of an infected female Phlebotomine sand fly. Leishmania are obligate intracellular protozoan parasites that infect mononuclear phagocytes2.Leishmaniasis is considered a serious public health problem on four continents: the Americas, Asia, and Africa. It has three general forms: cutaneous leishmaniasis (CL), mucocutaneous leishmaniasis (MCL), and visceral leishmaniasis (VL); these are caused by more than twenty species of 3, there are approximately one million cases worldwide, and the most common clinical manifestations comprise the appearance of painless single or multiple lesions in exposed areas, with well-defined borders. The mucocutaneous form has a different clinical manifestation involving marked destruction of mucosal regions, especially in the upper airways, that can provoke respiratory complications. Three to five percent of CL cases progress to the mucocutaneous form4.CL is the most prevalent clinical form. According to the World Health Organization5. The disease is fatal in over 95% of cases if not treated, and is considered the most severe form. It is endemic in 13 American countries, with an average of 3470 cases per year and a 7% fatality rate. In 2019, 97% of all cases were in Brazil7. The diagnosis is usually based on clinical criteria and specific tests, such as serology for the detection of antibodies and parasite isolation through bone marrow or splenic aspirates. In addition, the Kalazar Detect rK39 test is a rapid assay that can qualitatively determine the presence of antibodies against Leishmania8.The most common clinical manifestations of VL, also known as kala-azar, are fever, weight loss, hepatosplenomegaly, and hematological alterations9. Dayakar et al.10 suggest that the Th1/Th2 dichotomy plays a key role in VL. The development of more relevant clinical forms is associated with the predominance of Th2 cells, TGF-β, IL-4, and IL-10, which impair the antiparasitic activity of macrophages.The clinical forms of the leishmaniases are influenced by the host’s immune response. Upon parasitic infection, the innate immune response is triggered to restrain its proliferation. However, the disease progresses when there is suppression of reactive nitrogen and oxygen intermediates (RNI and ROI) within macrophages, responsible for inhibiting parasitic replication5. Nylén et al.11 reported that IL-10 in VL patient plasma enhances parasite replication in macrophages and that blocking IL-10 reduces parasite growth. Thus, IL-10 is associated with the more severe clinical manifestations because it inhibits the production of other mediators, such as IFN-γ and nitric oxide, and inhibits the leishmanicidal action of macrophages, preventing the elimination of intracellular parasites5. On the contrary, there are interleukins associated with controlling leishmaniasis, including IL-12, IL-17, IFN-γ, and TNF-α, that activate the acquired immune response13.Interleukin 10 (IL-10) is a regulatory cytokine produced by T cells, B cells, macrophages, dendritic cells (DCs), and epithelial cells14. These genetic factors are associated with distinct clinical manifestations of several diseases, and they generate important phenotypic differences in the population by coding proteins in different amounts or causing modification of the proteins themselves. Thus, the immune response can be induced in different ways depending on the polymorphism involved15.Single nucleotide polymorphisms (SNPs) are relevant in specific phenotypes that are more prone to develop diseases, considered risk factors17 several IL-10 polymorphisms have been investigated regarding their roles in leishmaniasis, but the results are inconsistent. Hence, the aim of the present study was to perform a systematic review and meta-analysis to determine if IL-10 SNPs influence the progression of leishmaniasis.Previous studies have shown that specific polymorphisms are associated with the regulatory role of IL-10 and disease progression18 and Preferred Reporting Items for Systematic Reviews and Meta-Analyses (PRISMA) guidelines19. This systematic review was registered in the PROSPERO database (CRD42021255374).This study was conducted following the Meta-analysis Of Observational Studies in Epidemiology (MOOSE) statementLeishmania-positive subjects with clinical manifestations of the disease, and the control group included asymptomatic individuals (Leishmania-positive) for VL and healthy individuals (Leishmania-negative) for CL. Studies were considered eligible if they satisfied the subsequent criteria: (i) case–control studies; (ii) subjects diagnosed with leishmaniasis by Montenegro skin test, myelogram, polymerase chain reaction (PCR), or serology tests, regardless of age and gender; (iii) SNP polymorphisms were investigated factors; (iv) control group composed of asymptomatic subjects as described above; and (v) sufficient information to evaluate the association between IL-10 polymorphism expression and disease progression. Family-based studies, editorials, comments and opinions, articles of reflection, projects, technical reports, reviews, and articles not related to IL-10 polymorphisms and leishmaniasis were excluded from this review.The present review focused on the subsequent question: Are the IL-10 polymorphisms risk factors for the progression of leishmaniasis? To evaluate the association between these variables, we assumed that all subjects were in an endemic area and allocated in groups (case or control) according to the stage of infection. The case group was composed of A systematic search using ScienceDirect, PubMed, and Scopus databases was performed to identify all relevant publications regarding the association of IL-10 SNPs in all types of leishmaniasis up to April 2021. The structured search strategy used the following terms: (“Interleukin 10” OR “IL-10”) AND (“polymorphism” OR “gene” OR “variant”) AND . To expand the number of eligible articles retrieved, no restrictions were imposed regarding study design or language.Two reviewers (R.R.S. and F.S.F.V.) independently screened the search results and identified studies that were potentially relevant based on the paper’s title and abstract. The selected articles were read in full and only the relevant ones, consistent with the eligibility criteria, were included in our study. Any disagreements were resolved after discussion with a third reviewer (D.S.T.). The reference lists of the retrieved studies were manually scanned to identify additional relevant studies.Data extraction from each manuscript was performed in consistence with the eligibility criteria, and the extracted data was recorded in a spreadsheet. The following were recorded from each article: name of the author(s), year of publication, total number of subjects (cases/controls), country where the research was conducted, leishmaniasis form, and IL-10 polymorphisms. We also listed genotype and allele counts for cases/controls and p values.20 by two independent reviewers (R.R.S. and F.S.F.V.). The NOS scale is made through a “star system” and judges the study based on three perspectives: selection of the study groups (maximum of one star for each column), the comparability of the groups (maximum of two stars), and the exposure (maximum of one star for each column). Studies that received a score of 6 or higher were considered as high quality.Studies that met the eligibility criteria were assessed for methodological quality using the Newcastle–Ottawa Scale (NOS)2 index21. When high heterogeneity (p < 0.1 for the Q test or I2 > 50%) was detected, the random-effects model was employed for meta-analysis; otherwise, a fixed-effects model was applied22.For articles that studied more than one polymorphism with different control groups, each result was considered individually for meta-analysis. Odds ratios (ORs) were used in meta-analysis to test the association between each SNP and the progression of leishmaniasis in case–control studies; 95% confidence intervals were included and p values < 0.05 were considered statistically significant. Five genetic models for each SNP were investigated corresponding to dominant, recessive, homozygote, heterozygote, and allelic modes of inheritance. Heterogeneity was evaluated using the chi-square-based Cochran’s Q statistic and the I23, publication bias was estimated by the qualitative effect of funding sources. Pearson’s chi-squared test was applied to examine Hardy–Weinberg equilibrium (HWE) in the healthy control group. The HWE was reached if p > 0.05. In all analyses, Review Manager Version 5.3 software  was used.Sensitivity analysis was performed by excluding individual studies from the pooled ORs and recalculating the statistical significance to explicate the stability of findings and to ascertain whether final pooled effect sizes were affected by a single publication. Since the number of included studies was less than 1030 were eligible according to the inclusion criteria and were included in this systematic review, but only five studies were subjected to statistical analysis  or full text (n = 2) evaluations. Finally, seven studies30, three in Brazil27, one in India29, and one in Turkey28. Table All eligible studies were published between 2013 and 2020. Two of the included studies were conducted in Iran24 and Covas et al.26, both analysing the rs180071 IL-10 SNP, indicated that IL-10 is not associated with CL and that the SNP does not regulate IL-10 secretion. Kirik et al.28 analysed the rs1800896 SNP and presented a deviation from HWE both in patient and control groups, and thus could not compare the risk estimate results. Both Hajilooi et al.30 studies demonstrated SNP influence in the progression of VL, particularly, rs1800871 and 1800896 IL-10 SNPs, respectively. Mishra et al.29 demonstrated that of the four SNPs analysed, only rs3024498 showed association with VL. However, Shehadeh et al.27 presented no association of IL-10 polymorphisms and VL progression, neither with rs1800871 nor rs1800896.Sonon et al.All studies included in this systematic review had an overall good methodological quality with NOS scores ranging from 6 to 8. All studies (100%) stated a research question with adequate case and control definitions, and clearly described information related to the eligibility criteria with the same method as for assessing the SNPs between the groups. A deficiency in the representativeness of the cases was seen, with 38% of the studies being without a star  and recessive models   and heterozygote models    and recessive models   and heterozygote models    and the deficit in immune response leading to the chronicity and severe manifestations of VL. This was more evident in heterozygous C/T patients than in homozygous patients, but there is not enough data to explain this difference26.This meta-analysis found that the SNP at the − 819 position did not show a relationship with poor prognosis in any genotype model, whether dominant, recessive, allelic, homozygous, or heterozygous, which supports other studies that have evaluated CL and concluded that the allele frequencies did not show remarkable differences between patients and controlsLeishmania, and geographic territory. This study did not show a relationship with any model and may emphasize that this polymorphism (rs1800896) cannot explain all forms of immune responses control in the patient. Innate and acquired immunity and other cytokines may also be involved in the pathogenesis30.Regarding the rs1800896 polymorphism studied, there was also no significant difference between patients and controls. Although the research has been carried out in several locations, since vulnerability to CL may not be related only to this SNP, but to several other aspects, such as ethnicity, species of 23, was investigated in a qualitative form, due to the insufficient number of primary studies in the literature. However, this is most likely not an issue herein since all studies had public funding and authors did not report conflicts of interest.Finally, it is important to consider the limitations of this study when analysing the results. There are few studies published on IL-10 polymorphisms and leishmaniasis, and most of them did not report any standardization on control groups. Moreover, small sample sizes and different methodological strategies, such as sample collection, ethnicity of subjects, forms of leishmaniasis, and sample processing of the included studies, may have affected the data analyses and the final heterogeneity results in some forests plots. Publication bias, which occurs when the authors support the interests of the private study’s financial sponsor, along with conflicts of interest, which may affect study design, conducting, and reporting (contributing to the omission of results or underreporting negative results29.Despite the evidence of the role of IL-10 in the persistence of leishmaniasis regardless of immune response, our findings demonstrate no significant association between rs1800871 and rs1800896 and the progression of leishmaniasis. Nevertheless, future research on the subject with greater methodological rigor and sample sizes is mandatory to clarify the role of IL-10 and its genetic profiles on disease progression and treatment


Autism spectrum disorders (ASDs) are one of the most common, highly heritable neurodevelopmental diseases affecting 1-2% of children under the age of 3. Although studies have implicated genetic predispositions, environmental risk factors, and maternal depression as the pathophysiology of ASD, it remains unclear. The association between antidepressant (AD) usage during pregnancy and the likelihood of ASD in children is still debatable. We carried out a systematic review to determine the relation of ASD with AD in offspring exposed to ADs in utero. We used the following terms of medical subject heading (MeSH) and keywords separately and in combination: “antidepressants,” “maternal/pregnancy depression,” “autism spectrum disorders/autism,” and “selective serotonin reuptake inhibitors (SSRI).” Our data search was conducted on PubMed, PubMed Central, Google Scholar, and Cochrane, which resulted in 28,141 articles. We identified and eliminated duplicates and then screened 9,965 articles by title and abstract. We then applied eligibility criteria over 143 relevant articles; a quality assessment was performed, and finally we included 18 selected studies.Mothers who had taken ADs during pregnancy for at least two medication prescription cycles and children detected to have ASD from two years to 18 years of age were included. We excluded articles in languages other than English, grey literature, case reports, letters to the editor, books, documents, animal studies, and studies published before 2017. Out of 18 studies, 17 evaluated ASD as the primary outcome, and for one study, the outcome was child behavioral as well as neurodevelopmental changes. Other additional outcomes studied were attention deficit hyperactivity disorder (ADHD), preterm birth, spontaneous abortion, small for gestational age, maternal mental illness, and persistent pulmonary hypertension. After adjusting for confounding factors, in six studies, the higher correlations between ASD and ADs were eliminated. Also, paternal AD use, maternal pre-conceptional AD drug use, and maternal depression itself are additional factors that raise the incidence of ASD. Autism spectrum disorder (ASD) is characterized by social impairment, poor communication, stereotypic behavior, aberrant sensitivity to sensory stimuli, and self-injurious conduct ,2. The pPathological findings in ASD include increased oxidative stress, hyperserotonemia, and loss of Purkinje cell integrity in the cerebellum . While tIn some circumstances, identifying modifiable environmental risk factors may benefit primary prevention. The prospect of uncovering one such modifiable causative factor has been raised by some reports suggesting that the usage of selective serotonin reuptake inhibitor (SSRI) antidepressants (ADs) during pregnancy could be related to ASDs in offspring. The importance of the serotonergic system in the pathophysiology of autism is gaining popularity, and prenatal exposure to serotonergic drugs is one biologically plausible avenue . BecauseADs are used to treat depression and anxiety disorders, which can be influenced by stressful life events . Up to 2According to animal research, in utero exposure to serotonergic ADs such as SSRIs may result in faulty serotonin signaling and loss of serotonin terminals in exposed fetuses, as well as changes in brain structure and function, which can lead to negative behavioral effects, culminating in autism phenotype ,7. PerinSome studies also suggest that psychiatric diseases share genetic determinants. Pregnant women who suffer from a mental illness during pregnancy for which ADs may be prescribed have a hereditary vulnerability that children may inherit. As a result, the higher risk of ASD in children of women who used ADs during pregnancy may be explained partially by genetic susceptibility rather than medication. Previous research on the relationship between maternal AD medication and ASD in offspring has been limited to investigations of ADs in general or SSRI ADs, resulting in limited information regarding the risk associated with a non-SSRI group of ADs ,14.Even though the majority of studies revealed evidence of correlation, conclusions varied due to concerns about confounding by indication. This was due to the possibility that depression or other psychiatric indications for AD use could be linked to autism via genetic or non-genetic pathways, and thus the possibility that the observed associations represent the risk of autism from the underlying maternal mental illness itself . ReportsMethodsSearch Strategy We followed the Preferred Reporting Items for Systematic reviews and Meta-Analyses (PRISMA) guidelines in our systematic review . ArticleInclusion CriteriaMothers who had taken ADs during pregnancy for at least two medication prescription cycles and children detected to have ASDs from 2 years to 18 years of age were included. And studies published from 2017 to 2022 were retrieved.Exclusion CriteriaWe excluded articles in languages other than English, grey literature, case reports, letters to the editor, books, documents, animal studies, and studies published before 2017.Risk of Bias in Individual StudiesTwo authors screened the retrieved articles by titles and abstracts and applied quality assessment tools. Studies with scores ranging from 7 to 8 were included in the study. We used the Newcastle-Ottawa bias scale tool for cohort studies, the assessment of multiple systematic reviews (AMSTAR) tool for systematic reviews, and meta-analysis and the scale for the quality assessment of narrative review articles (SANRA) checklist for reviews.ResultsStudy SelectionOur data search on PubMed, PubMed Central, Google Scholar, and Cochrane resulted in 28,141 articles. We identified and eliminated duplicates and then screened 9,965 articles by title and abstract. We then applied eligibility criteria and quality assessment tools to 143 relevant articles; we included 18 in the study. Out of 18 studies, 17 evaluated ASD as the primary outcome, and for one study, the outcome was child behavioral as well as neurodevelopmental changes. Two studies revealed no link between ADs and ASD, whereas eight studies found a positive association, out of which six studies revealed significantly reduced association on adjusting for controlling factors. Paternal AD use, periconceptional AD use, and maternal depression itself are additional factors that raise the incidence of ASD.DiscussionImpact of Depression During PregnancyAccording to epidemiologic studies, the prevalence of depression during pregnancy ranges from 8.5% to 11%, with up to 12.7% of women experiencing a major depressive episode. Prescriptions for ADs during pregnancy have increased considerably in the last 10 to 15 years, with estimates ranging from 4 to 16 times. According to research, untreated maternal depression, anxiety, and discontinuing ADs are associated with risks for both the mother and the growing child. This has been related to significant impairments in mother-child attachment and child development, as well as behavioral difficulties, low social engagement, and poor emotional regulation. As a result, physicians need to consider treating both the mother and the infant for prenatal depression .Serotonin-Induced Alterations in the Fetal BrainBecause of its efficacy, few side effects, and therapeutic safety, SSRIs are the most prescribed ADs. SSRI medications have been shown to easily pass across the placenta and blood-brain barrier. Neurogenesis, migration, and differentiation, among other elements of brain development, are all influenced by serotonin ,18. The Studies With Conflicting Results Regarding the Linkage Between AD and ASDZhou et al. presented an updated meta-analysis of the relation between maternal AD use during pregnancy and ASD. They included 14 studies, (eight cohort studies and six case-control studies). Pooled adjusted relative risk (RR) for cohort studies  was 1.13 (95% CI: 0.93-1.39), and pooled adjusted odds ratio (OR) for case-control studies  was 1.51 (95% CI: 1.15-1.99). The main bias was the limited number of studies. However, after adjusting to avoid publication bias, estimates demonstrated a non-significant association . Hence, the author does not recommend that pregnant woman who has been diagnosed with severe depression stop taking their ADs for fear of having an uncertain risk .From 2002 to 2010, Brown et al. conducted a retrospective cohort analysis of 35,906 singleton births in Ontario, Canada, comparing the risk of autistic spectrum disorder in children exposed to a serotonergic AD in utero to children who were not exposed. The average length of follow-up was 4.95 years, and the average age of the mothers was 26.7 years. The incidence of ASD was 4.51 per 1,000 person-years among children exposed to ADs versus 2.03 per 1,000 person-years among unexposed children (hazard ratio (HR): 2.16 with 95% CI: 1.64-2.86, adjusted HR: 1.59 with 95% CI: 1.17-2.17). The association was not significant  after controlling for confounding factors and when compared with unexposed siblings . One of the few issues with the study was that it only included women from Ontario's publicly funded lowest socioeconomic groups. Besides, outpatient data were restricted to diagnoses provided by a pediatrician or a psychiatrist to improve precision, and exposure status may have been misclassified if medications were filled but not utilized .A large Swedish cohort study by Rai et al. evaluated 254,610 individuals aged 4-17 years, including 5,378 with autism, living in Stockholm County from 2001 to 2011. Autism was diagnosed in 4.1% (n= 136) of the 3,342 children exposed to ADs during pregnancy compared to 2.9 % (n= 353) of the 12,325 children not exposed to ADs whose mothers had a history of psychiatric disease . The results of propensity score analysis and sibling control analysis were similar; however, only autism without an intellectual disability was linked to an increased risk in all of the studies. As a limitation, it was not possible to investigate trimester-specific or dosage response effects due to the limited sample size and the lack of comprehensive measures of depression severity assessment .ASD, attention deficit hyperactivity disorder (ADHD), preterm birth, persistent pulmonary hypertension, and spontaneous miscarriage were among the outcomes examined by Uguz and he included nine meta-analyses looking at the usage of any form of AD, and ASD was an outcome in 10 of the total studies. In AD exposure groups, the OR for the risk of autistic spectrum disorders was 1.22 to 1.81 for SSRIs, 1.13-1.81 for any AD, and 2.05 for non-SSRIs. Although this evaluation of current meta-analyses suggests that maternal AD usage is associated with a somewhat elevated risk of newborn and childhood outcomes, it is difficult to say if these effects are independent of underlying maternal psychiatric illnesses . Sujan eAn observational study by Ackerman et al. investigated prenatal AD exposure, maternal depression, the presence of an LGD (likely gene disrupted) mutation, and their interactions with ASD severity. The study included 2,748 patients with ASD, with an average age of 9. In conclusion, researchers discovered that AD exposure and the presence of an LGD mutation had a significant influence on ASD severity. A "two-hit" model is proposed, in which one variable establishes the basis for an initial risk that is then increased by a second variable. However, the small sample size and lack of information regarding out-of-study genetic variations were limitations of the study . In TablPotential Variables Affecting the OutcomeThe European Medicines Agency's (EMA) Pharmacovigilance Risk Assessment Committee (PRAC) concluded in September 2016 that the current evidence does not support a causal relationship and that the available studies on the risk of ASD after in utero SSRI exposure are contradictory, in part due to the different study designs and study populations chosen for analysis. Additionally, Morales et al. performed a systematic review of 15 observational studies  examining the risk of ASD associated with maternal AD exposure during pregnancy. Pooled effect estimates for the risk of ASD associated with maternal AD exposure during pregnancy compared to unexposed women with a history of affective disorder did not appear to be statistically significantly increased , whereas the risk of ASD associated with maternal AD exposure during the pre-pregnancy period compared to all unexposed women appeared statistically significantly elevated  .Another systematic review and meta-analysis by Mezzacappa et al. revealed a link between maternal AD usage and the likelihood of autistic spectrum disorder during pregnancy; however, the association appears to be more consistent throughout the preconception period than during each trimester . Six casYamamoto-Sasaki et al. conducted a retrospective cohort analysis to see if AD use was associated with ASD. From January 2005 to July 2014, they collected data on women and children, and children who were followed for 24 months after birth or until their withdrawal from the database. ADs were taken by 195 mothers of the 26,925 eligible mother-child pairs. The mean maternal age at birth of AD users and non-users was 32.1 and 31.4 years, respectively, and 81.5% of AD users were diagnosed with depression during pregnancy. According to a preliminary investigation, the prevalence of ASD is greater with any AD usage than without . However, statistical significance was lost when the analysis was controlled for the confounding effect of maternal depression during pregnancy . This finding backs up the theory that AD usage during pregnancy does not increase the risk of ASD in offspring. The paucity of socioeconomic data, the small number of women in our sample who used ADs during pregnancy, and the lack of data before the mother's pregnancy are all drawbacks [Another retrospective cohort study on the Swedish offspring 1, 580,629) by Sujan et al. revealed conflicting results. At the population level, first-trimester exposure was associated with an increased outcomes ratio  compared with unexposed offspring; however, after accounting for confounding factors, no increased risk of autism was noted in exposed children . The majority of ADs were SSRIs, and the study focused solely on the first trimester, which was considered a limitation of the study . Accordi by SujanDespite the discrepancies in the methodology used by the six meta-analyses, the conclusions were very consistent, according to Andrade. In general, AD use during pregnancy was linked to a higher risk of ASD in children. In studies that accounted for confounding variables and controlled for maternal mental illness; however, the magnitude and significance of the findings were reduced. ADs were also linked to a higher risk even when exposure was limited to the preconception period when the drugs could not have harmed the fetus. One of the most prominent limitations of meta-analyses is that they do not account for multiple hypothesis testing . Table 2Is a Group-Specific Study Required for Better Analysis?Kaplan et al. conducted a meta-analysis of four cohort studies to see if there was a link between prenatal SSRI exposure and ASD in children. The pooled effect size in the maternal psychiatric disorder with no SSRI exposure cohort ASD risk in children was somewhat larger than that of the SSRI exposure during the pregnancy cohort (OR: 1.81 versus 1.61 and 95% CI: 1.44-2.29 versus 1.16-2.25). The low number of studies included in the meta-analysis is the most significant drawback. Nonetheless, the inclusion of cohort studies, which are less prone to bias when measuring exposures during pregnancy, is set as a major strength of this study . HalvorsAraujo et al. reviewed 20 studies related to ASD, of which one study evaluated the effect of SSRI. The finding that prenatal exposure to SSRIs/serotonin-norepinephrine reuptake inhibitors (SNRIs) does not predict the risk of ASD is very compelling. It is worth noting that the findings of research strongly suggested that maternal depression, whether treated or not during pregnancy, raises the chance of ASD in the offspring . Leshem Another review by Rotem-Kohavi et al. focused on the neurobehavioral outcomes across childhood associated with prenatal SSRI exposure. They stated that maternal depression and autism may be linked through hereditary processes, making SSRI exposure a secondary risk factor underlying the effect of maternal mood . These findings also imply that both SSRI usage and maternal depression appear to influence an infant's long-term emotional and behavioral development, both alone and in combination. Inconsistencies in administrative data collecting and clinical classifications of both ASD and maternal depressive mood symptoms hampered this research. The fact that depression is a chronic illness, as well as the number of episodes that occur during the prenatal and postnatal years, adds to the confusion . Table 3LimitationsOur systematic review has a few limitations. We selected articles published in English and collected data from the last five years; therefore, additional supporting data might have been missed. Lack of the more complete covariate  association studies, along with conflicting results of some studies are other limitations of this review. There is a need for comprehensive studies with a bigger sample size and including more covariates to obtain a detailed analysis.Our systematic evaluation validates the relationship between AD exposure in utero and the risk of ASD, but it raises strong reservations regarding causality because there is no substantial risk increase when the SSRI exposure and non-exposure groups are adjusted for covariates. In literature, maternal serotonin is linked to fetal neurodevelopmental changes; hence, children exposed to AD medicines during pregnancy may have a little higher chance of having impaired motor development. In addition, any incremental risk must be weighed against the negative repercussions of inadequately treating depression and related disorders. These factors must be considered when deciding whether to use ADs during pregnancy, considering the specific patient and family circumstances. Additional research is needed, particularly regarding maternal psychiatric disorders, periconceptional AD use, and paternal AD usage.


There is keen interest in better understanding the impacts of alpha-linolenic acid (ALA), a plant-derived n-3 fatty acid, in ameliorating the development of cancer; however, results of several prospective cohorts present an inconsistent association between ALA intake and the incident colorectal cancer (CRC). We aimed to investigate the summary association of dietary intake and biomarkers of ALA with CRC risk based on the prospective cohorts.Pertinent prospective cohorts were identified in Cochrane Library, PubMed, and EMBASE from inception to February 2022. Study-specific risk ratios (RRs) with 95% confidence intervals (CIs) for comparing the top with the bottom quartiles of ALA levels were combined using a random-effects model. Nonlinear dose-response relationships of ALA levels in diet and blood with CRC risk were assessed using the restricted cubic spline models, respectively.2: 0.00%) for dietary intake and 0.83  for biomarker. Each 0.1% increase in the levels of ALA in blood was associated with a 10% reduction in risk of CRC , whereas no significant dose-response association was found between dietary intake of ALA and the incident CRC .Over the duration of follow-up with a median of 9.3 years ranging from 1 to 28 years, 12,239 CRC cases occurred among 861,725 participants from 15 cohorts (11 studies on diet and 5 studies on biomarkers including 4 on blood and 1 on adipose tissue). The summary RR was 1.03 (95% CI: 0.97, 1.10; IBlood levels of ALA were inversely and linearly associated with the risk of CRC, which suggested that increased intake of ALA to improve circulating levels was beneficial for CRC prevention. Colorectal cancer (CRC) is the second most common cancer diagnosed in women and the third most in men, and currently ranks as the fourth most deadly cancer worldwide with nearly 900,000 deaths annually . The incDietary factors have been shown to play an important role in the prevention of CRC . At cellin vivo  with CRC risk using the available data from the more comprehensive perspective studies.We identified 20,195 potential studies from PubMed, EMBASE, and Cochrane Library databases up through Feb 2022, and the search strategy we have predefined was listed in the literature searching section of To assess the association of ALA in diet and human tissues and the risk of CRC, the inclusion criteria were: (1) Participants: Adults of any age across different countries; (2) Exposure: levels of ALA intake estimated by dietary records, food frequency questionnaires, or quantitative determining the compositions or concentrations of ALA in circulating blood and adipose tissue (AT); (3) Outcomes: Evaluating the incident CRC as an endpoint, presented as multivariate-adjusted risk ratio (RR) or hazard ration (HR) with 95% confidence interval (CI); (4) Study design: Prospective cohort study, nested case-control study, and case-cohort study.The following data were extracted by the two independent reviewers from each original study using a standardized extraction form: first author, publication year, study design (prospective cohort/nested case-control/case cohort), study location (America/Europe/Asia), cohort name, sample size (number of cases/participants), baseline age , gender, duration of follow-up , cancer location (colon/rectum), exposure measurements, types of interest exposure (diet or biomarker), multivariate-adjusted RRs (HRs) with 95% CI for all category levels of ALA in diet or human tissues, and the potential confounders adjusted. The study quality of each included study was evaluated by using the 9-stars Newcastle-Ottawa Scale (NOS)  Supplem.If an original study provided HRs with 95% CIs for the incident CRC, the HR value was assumed to approximate the RR value. All the included studies provided RR (HR) for ALA intake (diet or biomarker) based on various categories  or per SD difference in exposure. To achieve a consistent approach to the present meta-analysis, the RRs (HRs) were first transformed to involve comparisons between the top and the bottom quartiles of baseline diet or biomarker of ALA using methods described previously , 22. In 2 statistic , baseline age , gender (man vs. women), median duration of follow-up , cancer location (colon vs. rectum), quality scores (7 vs. 8–9), study design (prospective cohort vs. nested case-control/case-cohort), biomarker types (adipose vs. blood), and multiple adjustments (yes vs. no). A univariate meta-regression with restricted maximum likelihood was performed to measure if summary RR significantly differed between each stratum analyzed. Sensitivity analyses were performed to evaluate the possible influence of individual studies on the summary results. A possibility of publication bias was qualitatively delineated by the asymmetry of funnel plots and quantitatively evaluated by Egger's regression tests (Multivariate-adjusted RRs (HRs) comparing the top with the bottom quartiles of ALA intake (diet and biomarker) in each study were first transformed to their logarithm (logRRs), and their corresponding 95% CIs were used to calculate the standard errors (selogRRs). Summary RRs (SRRs) with 95% CIs as the overall risk estimate for the top vs. bottom quartiles of ALA intake was calculated using a random-effects model described by DerSimonian and Laird , which ctatistic . We defion tests .P-value for non-linearity was calculated to detect a potential departure from a simpler linear trend by testing the coefficient of the second spline equal to zero .Dose-response meta-analyses were conducted to determine whether the levels of ALA in diet or circulating blood were dose-dependently associated with the risk of CRC. In brief, individual studies with three or more categories were included in the dose-response analysis and the median values of ALA levels in both diet and blood for each exposure category were assigned as previously described . A curvi to zero . In the  to zero . Two-taiThe major result of the search strategy is presented in the PRISMA flow diagram . The iniThe characteristics of 15 independent prospective studies are presented in 2 = 0.00%). An inverse association was found between ALA biomarker and CRC , with no between-study heterogeneity (I2 = 0.00%).The pooled association comparing the top with the bottom quartiles of dietary intake and biomarkers of ALA were presented in In the stratified analysis of dietary ALA intake concerning CRC , there wIn sensitivity analyses that exclude one study at a time and reanalyzed the remaining data, the exclusion of any individual study of ALA in the diet as interest exposure did not substantially change the summary result . As for In publication bias analyses for either dietary intake or biomarker, no publication bias was indicated by Begg's funnel plot  , 4 or EgNine cohorts with dietary intake of ALA were available for the dose-response analyses –38. Ther2 = 35.9%)  .To the best of our knowledge, this present study is the first meta-analysis that specially focused on the impacts of plant-based n-3 ALA (diet vs. biomarker) on the risk of CRC. Our pooled analysis of prospective cohorts suggested that blood levels of ALA were linearly and inversely associated with CRC risk, but no significant association was found for dietary intake of ALA. Such findings support that increased levels of ALA intake have potential benefits in preventing the development of CRC, which may further extend the previous meta-analyses with mixed n-3 PUFAs as interest exposure to highlight that the plant-derived ALA remains a protective nutrient for the incident CRC , 10, 15.Results of our meta-analysis based on the prospective cohorts with the dietary estimation of AL showed a null association with CRC risk, and there was no significant difference by age, gender, geographical regions, cancer locations, duration of follow-up, or multiple adjustments. Compared with our present study, most of the previous studies especially focused on food n-3 PUFAs mixed plant- with marine-based sources , 10. ThePUFA levels in human tissue  are currently regarded as a reasonable biological marker of habitual dietary fat intake, with sufficient evidence in the strong correlation between dietary fatty acid intake and circulating levels even if it is a single blood sample. Results of our meta-analysis based on five prospective cohorts revealed that increased levels of ALA in biospecimen (blood and AT) were significantly associated with a reduced risk of CRC. In support of these major findings, a similar inverse association with biomarker ALA was also observed in two publications of population-based epidemiological studies , 17. NevIn the stratified analysis of biomarker ALA in relation to CRC, we found a beneficial association estimation in males rather than in females. Given that estrogen might have participated in the etiology of CRC , 43, losin vivo, which can be progressively transferred to eicosapentaenoic acid (EPA), docosapentaenoic acid (DPA), and finally docosahexaenoic acid (DHA)  , throughid (DHA) . ALA migid (DHA) . In addiid (DHA) , 51. Thiid (DHA) . Finallyid (DHA) , 53, 54.Several strengths are currently emphasized in our study. The eligible prospective cohort studies only were included, and therefore no recall and selection bias caused by retrospective studies would influence the summary result. Besides, stratified analyses with a meta-regression test indicated that the overall association estimations were not affected by the strata analyzed such as age, follow-up years, cancer location, and multiple adjustments, thereby increasing the potential possibility of the robust performance of final results. Third, compared with previous meta-analyses , 15, we There are also several limitations in the present study. First, sensitivity analysis for biomarkers ALA indicated that exclusion of one study would potentially change the direction of the overall result toward a null (18). However, this study enrolled volunteers from a selected population of highly educated women, which was not representative of the general population. We therefore cannot rule out the possibility that selection bias might have seriously affected the association estimated in this study. Second, RRs (HRs) for various category levels in each original study were transformed with the top vs. bottom quartiles to provide a consistent approach to the meta-analysis, in which systematic error might have occurred during the data transformation. Third, though each original study controlled multiple confounding factors, there were still some residual confounders that might have changed the direction of the summary association. Fourth, though the beneficial association for the biomarker of ALA was found to be more significant in male and middle-aged populations, the results based on the subgroup analyses may not be popularized because of the limited number of included studies. Fifth, dietary changes or changes in food compositions may have occurred after blood collections and before the onset of CRC, perhaps leading to an underestimation of the pooled association. Sixth, misclassification in dietary estimations is inevitable, which was likely to bias the pooled association toward a null. Finally, we found a significant inverse association for ALA levels in the blood, but the results of AT measurement in relation to CRC risk are needed to be interpreted with more caution because of the limited number of published cohorts.The current meta-analysis indicated that biomarkers of ALA were inversely associated with the incident CRC, and each 0.1% increase in circulating levels of ALA was associated with 10% reduction in CRC risk. Encouraging the consumption of foods rich in ALA to improve its levels in the blood may potentially decrease the risk of CRC. Nevertheless, well-designed and large-scale cohorts with biomarkers are still needed for better reconfirming the potential impacts of ALA intake in the primary prevention of CRC.The original contributions presented in the study are included in the article/BY conceived the idea and designed the study strategy and provided critical revisions of the manuscript for important intellectual content, administrative and funding support, and supervision. Z-BD, Y-LX, YT, and B-BH conducted a reference search. Z-BD summarized the data and conducted data acquisition and statistical analyses. Z-BD and X-LR drafted the manuscript. All authors contributed to the article and approved the submitted version.This work is supported by the Programs Foundation of the Wenzhou Medical University of Zhejiang Province, China (89217015), the Scientific Technician Funding of Wenzhou Science and Technology Bureau (X20210055), the Basic Medicine Funding of Wenzhou Science and Technology Bureau (Y20180195), and by the 2017 Chinese Nutrition Society (CNS) Nutrition Research Foundation—DSM Research Fund (95017008). The funders have no role in study design, data collection and analysis, decision to publish, or preparation of the 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.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.


Angiotensin-converting enzyme inhibitors (ACEIs) and angiotensin receptor blockers (ARBs) are widely used in the treatment of hypertension. Hypertension is often accompanied by osteoporosis. However, the relationship between ACEI/ARB and fractures remains controversial. The purpose of this meta-analysis was to update the potential relationship between ACEI/ARB and fractures.  This meta-analysis was identified through PubMed, EMBASE, Cochrane Library, and Web of Science. Related studies about ACEI/ARB with the risk of fracture were published from inception to June 2022. P < 0.001) for composite fractures and 0.96  for hip fractures; the RRs of ARB compared to the nonusers were 0.82  for composite fractures and 0.85  for hip fractures. Furthermore, in the subgroup analysis, male may benefit from ARB , and the European may also benefit from ARB .  Nine qualified prospective designed studies, involving 3,649,785 subjects, were included in this analysis. Overall, the RRs of ACEI compared with the nonusers were 0.98 (95% CI: 0.88, 1.10;  ACEI usage will not decrease the risk of osteoporosis fracture. On the contrary, ARB usage can decrease the risk of total fracture and hip fracture, especially for males and Europeans. Compared with ACEI, for patients at higher risk of fracture in cardiovascular diseases such as hypertension, the protective effect of ARB should be considered. With worldwide growth in aging populations, the number of people suffering from osteoporosis is also increasing and brings a battery of physical and mental pain and enormous economic burdens to the individual and society. Fracture, especially hip fracture, is a serious complication of osteoporosis. Hip fractures often result in paralysis and loss of self-care, and there was a twenty percent increase in the mortality rate during the first year . At presThe renin-angiotensin-aldosterone system (RAAS) is a pressure-boosting regulatory system produced by the kidneys in the body that can produce angiotensin II (Ang II), thereby raising blood pressure. RAAS blockers have been widely used in patients with hypertension  and mainThe relationship between RAAS blockers and bone health, structure, and metabolism has received increasing attention, but the answer remains uncertain yet –10. AbuoThe literature search for articles published from inception to June 2022 was conducted in MEDLINE (via PubMed), EMBASE, Cochrane Library, and Web of Science. Mesh and free text terms used for the search were combined with methodological filters. All searches were limited to the English language, and studies were conducted with healthy humans aged 18 years.The following search terms were used: “angiotensin-converting enzyme inhibitors,” “angiotensin II receptor blocker,” “angiotensin receptor blockers,” “inhibitors, ACE,” “ACEI,” “benazepril,” “captopril,” “enalapril,” “ARB,” “valsartan,” “irbesartan,” “losartan”; AND: “fracture,” “osteoporosis,” “fall injuries,” “bone mineral density,” “BMD,” “bone loss”, odds ratios (ORs), hazard ratios (HRs), or other measures that could be transformed into RRs, with 95% confidence intervals (CIs);Only cohort studies were considered.All studies included were in accordance with PICO principles, and five eligibility criteria were as follows:If different papers came from the same cohort, the paper with the most comprehensive design was brought into this analysis. Those citations both reviewers deemed irrelevant were excluded. Citations with disagreement were also included for a full review.As shown in Supplement Two reviewers independently used data extraction tables to extract data from various included studies. The following items were evaluated and extracted: author and year of publication; methods ; participants ; intervention type ; fracture site; and the effect size. For subgroup analysis, we also stratified the data according to restriction factors, such as fracture site, sample size, location, gender, and quality level.Q statistic and the I2 index [P value of 0.05 or less.RRs were used as a common measure of the association between ACEI/ARB use and fracture risk . In thisI2 index . Use forA total of 2916 potentially relevant citations were obtained through the all-sided search. As a result, we included 9 cohort studies , 20–25. The studies were from different regions: three from America , 20, 25,I2 = 93.5%, P < 0.001). However, the risk of composite fractures was significantly reduced in subjects receiving ARB compared to nonusers .The RR (95% CI) of ACEI/ARB usage associated with composite fractures is summarized in I2 = 62.1%, P=0.048). In contrast, the usage of ARB would reduce the risk of hip fracture as compared to nonusers .The effects of ACEI/ARB on the risk of hip fracture are summarized in I2 = 71.9%, P=0.028). What's more, when grouped by regions, the European may benefit from ARB blockers . However, in the female subgroup, the ACEI/ARB usage was not statistically significant. ACEI usage did not have a positive effect on fracture in any of the groups. In the grouping of sample size and quality level, both the combined results of five large-sample size cohort studies [2 = 87.5%, P < 0.001).The relative risk of fracture associated with ACEI/ARB usage by subgroups is summarized in Supplement  studies , 20–23 a studies , 23, 25  studies , 23 werePublication bias is initially examined by drawing funnel plots. Then, check it further with Begg's test and Egger's test. For ACEI, ARB and ACEI or ARB use, the Begg's test for publication bias was not significant . For ACEI, ARB and ACEI or ARB use, the Egger's test was also not significant . Additionally, the use of the trim and fill correction procedure did not alter the results. No evidence of publication bias was detected.We conducted a comprehensive meta-analysis, including data from 9 qualified prospective cohort studies investigating the association between ACEI/ARB use and the risk of fracture. We found that ACEI use was not significantly associated with fracture risk, but ARB use was associated with a lower risk of fracture. Besides, by the subgroup analysis, males and European may benefit from ARB.Our updated meta-analysis differs in several important ways from the previous review performed by Kunutsor et al. . First, The mechanisms underlying the beneficial effects of RAAS inhibitors on bone fracture remain debated in recent years. Apart from the more well-known systemic RAAS activations, the tissue RAAS activation also plays an important role via the endocrine effects, which mediate important physiological stimuli including bone and osteoporosis . A growiSome mechanisms may explain the difference between ACEI and ARB. First, the activation of RASS can increase the plasma concentration level of Ang II, which leads to a series of physiological effects. Although ACEI and ARB share the same mechanism pathway , neither ACEI nor ARB can completely inhibit the production of Ang II and the toxic effects of Ang II . The relThe subgroup analysis has shown that men would benefit from the ARB use, but it was not statistically significant for women. This may be because women's bone metabolism and bone density are more affected by estrogen than ARB , which hWe minimize the sources of heterogeneity by strictly implementing inclusion and exclusion criteria and excluding low-quality studies. A random-effect model was used to eliminate some heterogeneity. Subgroup analysis and sensitivity analysis were used to explore the source of heterogeneity. There was significant heterogeneity in this study, which may be due to different dosage, regimen, duration, and other reasons. In addition, bone metabolism is also affected by the gender and age of the population, and postmenopausal women and elderly people are more prone to fracture, so the ratio of gender and age will also lead to the generation of heterogeneity. Heterogeneity may be caused by whether the subjects take drugs that affect bone metabolism such as vitamin D and calcium tablets, as well as unclear conditions such as sun exposure and exercise.This study also has several limitations. First, many of the studies suffer from significant sources of bias, and the effect in many occasions was assessed by very few studies. Second, due to the limited information we gathered from the original studies, we were unable to obtain treatment time and dose data for further analysis. Third, none of the included studies mentioned bone metabolism indexes and bone mineral density, so these aspects were not analyzed. Fourth, several drugs are often used concurrently in clinical, making it challenging to assess specific drug effects. So more high-quality studies are needed.In conclusion, this meta-analysis demonstrated that ACEI cannot decelerate the risk of fracture, whereas ARB has a protective effect on fracture risk. Hence, in clinical practice, ARB may be a priority when patients suffer from hypertension with a higher risk of fracture.


I2 was used to evaluate heterogeneity among studies. A total of 72 eligible articles were included in the meta-analysis. It was found that the ferritin levels of patients increased with the severity of the disease, whereas their serum iron levels and hemoglobin levels showed opposite trends. In addition, non-survivors had higher ferritin levels (SMD (95%CI): 1.121 ; Z = 8.22 p for Z < 0.001; I2 = 95.7%, p for I2 < 0.001), lower serum iron levels (SMD (95%CI): −0.483 , Z = 8.27, p for Z < 0.001; I2 = 0.9%, p for I2 =0.423) and significantly lower TIBC levels (SMD (95%CI): −0.612 , Z = 4.16, p for Z < 0.001; I2 = 71%, p for I2 = 0.016) than survivors. This meta-analysis demonstrates that ferritin, serum iron, hemoglobin and total iron banding capacity (TIBC) levels are strongly associated with the risk, severity and mortality of COVID-19, providing strong evidence for their potential in predicting disease occurrence and progression.The coronavirus disease 2019 (COVID-19), caused by severe acute respiratory syndrome coronavirus 2 (SARS-CoV-2), is spreading rapidly around the world and has led to millions of infections and deaths. Growing evidence indicates that iron metabolism is associated with COVID-19 progression, and iron-related biomarkers have great potential for detecting these diseases. However, the results of previous studies are conflicting, and there is not consistent numerical magnitude relationship between those biomarkers and COVID-19. Thereby, we aimed to integrate the results of current studies and to further explore their relationships through a meta-analysis. We searched peer-reviewed literature in PubMed, Scopus and Web of Science up to 31 May 2022. A random effects model was used for pooling standard mean difference (SMD) and the calculation of the corresponding 95% confidence interval (CI).  Coronavirus disease 2019 (COVID-19), caused by severe acute respiratory syndrome coronavirus 2 (SARS-CoV-2) . The rapThe clinical manifestations of COVID-19 are various, from asymptomatic infection to death. Therefore, it is very necessary to find a reliable early biomarker to identify the emergence and progression of disease ,4. NumerThough serum or plasma iron is a main indicator for iron homeostasis, ferritin, hemoglobin, hepcidin, TIBC and transferrin saturation (TSAT) would explain iron distribution more roundly. Previous studies have investigated the relationships among iron, ferritin, hemoglobin, hepcidin, TIBC, TSAT and COVID-19, but their results are conflicting and inconsistent. Most studies about serum iron have shown that lower iron levels are found in more at-risk groups, such as non-survivors compared to survivors ,11,12,13Considering that any individual study may not have the sufficient power to obtain a reliable conclusion, this meta-analysis was conducted to: (1) Summarize and evaluate the results of numerous papers on the relationships between serum iron, ferritin, hemoglobin, TIBC, TSAT and hepcidin levels and the mortality and clinical severity of SARS-CoV-2 patients. (2) Assess the potential between-study heterogeneity; and eventually investigate the potential publication bias.Our meta-analysis searched peer-reviewed literature on PubMed , Web of Articles were included based on the following four criteria: (1) the cases had COVID-19; (2) observational study design; (3) samples were from blood; (4) data were reported as mean ± standard deviation (SD) or the other forms could be translated to mean ± SD. Some studies were excluded if they were: (1) experimental studies (2) reviews.If the data were duplicated in one study, only the one that involved the largest number of cases was included. Two investigators searched for the articles and extracted data independently. The disagreements about eligibility of an article were settled by discussion.On each article, the following information and data were recorded: the first author, county (continent), publication year, study design, disease assessment, outcome groups, gender, mean age, number of cases, number of controls, number of non-survivors, number of survivors, numbers of severe and non-severe cases, mean ± SD of these groups and data units. If just standard error mean (SEM) was available, SD was calculated by the formula SEM = SD/ website .I2 was used to assess the heterogeneity among studies. For the purpose of pooling effect sizes and all the analyses, we used a random effects model. All reported probabilities  were two-sided, and p < 0.05 was recognized as statistically significant. If meta-regression demonstrated sources of heterogeneity in continent, publication year, study design and subgroup analyses were performed.The meta-analysis was performed using the software Stata 15.0 . In order to estimate the association between serum-iron-related biomarkers and the onset and progression of COVID-19, we used the standard mean difference (SMD) with the corresponding 95% confidence interval (CI) as the effect size. The SMD is the ratio of the mean difference to the pooled standard deviation. After the preliminary search, 6916 articles were identified, 2040 from PubMed, 1390 from Web of Science and 3486 from Scopus . After sp for Z < 0.001; I2 = 95.7%, p for I2 < 0.001; A total of twenty-nine studies assessed the connections between ferritin levels and the mortality of SARS-CoV-2 patients in this meta-analysis, involving 2131 non-survivors and 7813 survivors. The ferritin levels were significantly higher in the dead patients than that in survivors (SMD (95%CI): 1.121 ; Z = 8.22 p for Z < 0.001; I2 = 96.3%, p for I2 < 0.001; Seventeen studies about ferritin were included in the COVID-19 risk analysis, including 964 cases and 966 controls. The overall comparison demonstrated that cases had higher ferritin levels than controls (SMD (95%CI): 1.383 ; Z = 4.58 p for Z < 0.001; I2 = 96.00%, p for I2 < 0.001; p for Z < 0.001; I2 = 92.7%, p for I2 < 0.001; p for Z = 0.003; I2 = 97.2%, p for I2 < 0.001; In terms of the relationship between ferritin and the severity of COVID-19, a total of thirty-nine studies were included, involving 1681 severe, 1337 non-severe, 433 moderate and 1416 mild individuals. The results reported that the ferritin levels of severe groups were higher than those of the non-severe groups (SMD (95%CI): 0.864 ; Z = 3.57 p for Z < 0.001; I2 = 0.9%, p for I2 =0.423; A total of seven studies included analyses of serum iron levels and the mortality of SARS-CoV-2 patients, involving 456 non-survivors and 1508 survivors. The serum iron level in dead patients was significantly lower than that in alive patients (SMD (95%CI): −0.483 , Z = 8.27, p for Z = 0.001; I2 = 96.7%, p for I2 < 0.001; Seven studies were included in the serum iron and COVID-19 risk meta-analysis, including 492 cases and 704 controls. p for Z = 0.033; I2 = 0.00%, p for I2 =0.545; p for Z = 0.014; I2 = 94.2%, p for I2 < 0.001; A total of ten studies were included in the analysis of the relationship between the severity of COVID-19 and serum iron, involving 298 severe, 113 mild and 340 non-severe patients. Severe groups had lower iron levels than the mild groups (SMD (95%CI): −0.293 ; Z = 2.13 p for Z = 0.343; I2 = 82.5%, p for I2 < 0.001; Regarding the connection between hemoglobin level and the mortality of COVID-19, a total of six studies were included, involving 276 the deceased and 1029 survivors. However, no significant differences were found between the hemoglobin levels of dead and recovered patients (SMD (95%CI): −0.186 , Z = 0.95, p for Z = 0.028; I2 = 87.9%, p for I2 < 0.001; Seven studies were included in the hemoglobin level and COVID-19 risk meta-analysis, involving 367 cases and 511 controls. p for Z = 0.298; I2 = 5.80%, p for I2 =0.386; p for Z = 0.012; I2 = 86.50%, p for I2 < 0.001; In terms of the analysis of severity, a total of twenty studies were included, including 1060 severe, 524 mild and 851 non-severe individuals. Though the difference between severe and mild groups (SMD (95%CI): −0.073 ; Z = 1.040 p for Z = 0.232; I2 = 84.8%, p for I2 =0.001; A total of three studies on the relationships between hepcidin and the mortality of COVID-19 were included in this meta-analysis, including 66 non-survivors and 224 survivors. Dead patients had slightly higher hepcidin levels than recovered patients, but this was not significant (SMD (95%CI): 0.447 ; Z = 1.190 p for Z = 0.345; I2 = 96.40%, p for I2 < 0.001; The analysis of COVID-19 risk included four studies totally, with 177 cases and 261 controls. Though higher hepcidin level was found in cases than that in controls, the difference was not significant (SMD (95%CI): 0.750 ; Z = 0.95 p for Z < 0.001; I2 = 71%, p for I2 = 0.016; p for Z = 0.521; I2 = 59.6%, p for I2 = 0.084; Four studies about TIBC were included in the analyses of mortality, including 382 non-survivors and 719 survivors. Analyses about TSAT involved three studies, 281 non-survivors and 458 survivors totally. The TIBC level was significantly lower in the death group than in the survivors group (SMD (95%CI): −0.612 , Z = 4.16, Strong evidence of heterogeneity among studies was documented for the relationships between these iron-related biomarkers and mortality, clinical severity or risk in SARS-CoV-2 patients. The heterogeneity for the between-study was explored through the univariate meta-regression with the covariates of continent, gender, study types, years, ages and numbers of samples in the analysis. Meta-regression indicated that 75.02% of the heterogeneity in the COVID-19 risk analysis in serum iron was explained by continent. In terms of ferritin, publication year and sample size of the death group explained 18.33% and 14.78% of the heterogeneity of the mortality analysis, respectively; study type contributed 21.04% of the heterogeneity of overall analysis; and publication year contributed 20.33% of the heterogeneity of the severe–mild comparison. In addition, publication year and continent explained 46.95% and 46.09% of the heterogeneity of the moderate–mild analysis in ferritin, respectively. Regarding the heterogeneity of hemoglobin and mortality analysis, publication year accounted for 69.79%.I2 > 50% as the criterion to evaluate the robustness of conclusion of ferritin. Twelve studies [p < 0.001; I2 = 44.2%) and risk  analyses of ferritin.Considering the potential small-study effects, Galbraith plot showed the contribution of results from the different relevant studies to the heterogeneity. Then the “leave- one-out” sensitivity analyses used  studies ,59,60,62 studies ,20,21,24p = 0.930) or risk analysis (p = 0.129) of serum iron. However, the risk (p = 0.009) and mortality analyses (p < 0.001) of ferritin had significant publication bias. In terms of hemoglobin, Egger’s test demonstrated no publication bias in the analyses of mortality (p = 0.700) and risk (p = 0.483). Publication bias in mortality analyses of TIBC (p = 0.077) and TSAT (p = 0.358) was also not found. Moreover, no publication bias was found in the mortality (p = 0.065) and risk analyses (p = 0.699) of hepcidin.The results of sensitivity analysis indicated that no individual study had an excessive influence on the pooled measure for all comparisons . Egger’sOur meta-analysis was based on 72 articles, containing 148 studies, of which 85 were about ferritin, 22 were about serum iron, 27 were about hemoglobin, 7 were about hepcidin, 3 were about TSAT and 4 were about TIBC. Mortality analyses included 29 studies on ferritin, 8 on serum iron, 6 on hemoglobin, 3 on hepcidin, 4 on TIBC and 3 on TSAT. Our results indicated that the serum iron and TIBC levels of the deceased were significantly lower than those of the survivors, but higher ferritin levels were found in the deceased. In terms of hemoglobin, hepcidin and TSAT, we did not find an association with death outcome. The risk analyses included 17 articles on ferritin, 7 on serum iron, 7 on hemoglobin and 4 on hepcidin. It was shown that cases had lower serum iron and hemoglobin levels, but higher ferritin levels than controls. There were not significant results in hepcidin analyses. Regarding the analyses of clinical severity, 39 studies on ferritin, 14 on hemoglobin and 7 on serum iron were included. All of them were found to be related to the severity of COVID-19.According to previous studies, excessive inflammation is a characteristic of COVID-19 . Iron plHepcidin is a main regulatory factor of iron metabolism that is associated with body iron level ,84. HoweFerritin is also recognized as an acute phase reactant of inflammation, influenced by the presence of iron, hepcidin ,88,89,90In meta-analysis, between-study heterogeneity is common. Thus, exploring the sources of between-study heterogeneity is essential. We performed univariate meta-regression, with covariables such as continent, study type, publication year and sample size. The regression results explained part of the heterogeneity in our meta-analysis, but there was still some heterogeneity not being detected. In addition, the Galbraith analysis indicated that high heterogeneity in ferritin resulted from twelve mortality analysis studies and five risk analysis studies. After excluding these studies, low heterogeneity and robust results without small-study effect were documented. However, the final results of ferritin were not changed.There were some strengths in our meta-analysis: First of all, as far as we know, our study has unified a large number of studies on the associations of iron-related biomarkers with risk, clinical severity and mortality in COVID-19 patients, avoiding inaccurate conclusions of individual studies. More importantly, the random effects were used to estimate the pooled SMD. Thus, it was still possible to draw convincing results though the inconsistent measurement conditions and units for iron-related biomarkers in different studies.However, our study has several limitations. Firstly, high heterogeneity was found in almost all indicator analyses, but some of analytical results were not explained by meta-regression or subgroup analysis. We cannot get a more accurate evaluation for the sources of the heterogeneity due to the lack of corresponding study-level covariates in the reported articles. Moreover, the differences in iron-related biomarkers’ reference values between females and males illustrated the importance of gender for our study. However, we could not conduct meta-analysis including gender for the reason that there was not enough information about it.Our meta-analysis showed that the levels of serum iron and TIBC in dead patients were significantly lower than in survivors, and the ferritin level was higher in death groups than in survivors, whereas the relationship between hemoglobin and mortality was not significant. Moreover, serum iron and hemoglobin levels were lower in cases and negatively correlated with the severity; on the contrary, ferritin level was higher in cases. In addition, no statistically significant results were found in the hepcidin and TSAT levels of the severity and mortality groups. That was possibly due to the limited number of studies.In conclusion, we found that ferritin, serum iron, hemoglobin and TIBC levels are closely associated with the risk, severity or mortality of COVID-19. These results provide strong evidence for the applications of iron-related biomarkers in the prediction of the COVID-19 occurrence and development. Moreover, lower serum iron and hemoglobin levels could provide clues for explaining the deteriorated process of COVID-19. However, future studies are needed to further confirm these results in future research.