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Modern societies are experiencing an increasing trend of reduced sleep duration, with nocturnal sleeping time below the recommended ranges for health. Epidemiological and laboratory studies have demonstrated detrimental effects of sleep deprivation on health. Sleep exerts an immune-supportive function, promoting host defense against infection and inflammatory insults. Sleep deprivation has been associated with alterations of innate and adaptive immune parameters, leading to a chronic inflammatory state and an increased risk for infectious/inflammatory pathologies, including cardiometabolic, neoplastic, autoimmune and neurodegenerative diseases. Here, we review recent advancements on the immune responses to sleep deprivation as evidenced by experimental and epidemiological studies, the pathophysiology, and the role for the sleep deprivation-induced immune changes in increasing the risk for chronic diseases. Gaps in knowledge and methodological pitfalls still remain. Further understanding of the causal relationship between sleep deprivation and immune deregulation would help to identify individuals at risk for disease and to prevent adverse health outcomes.

Epidemiological evidence indicates that regular physical activity and/or frequent structured exercise reduces the incidence of many chronic diseases in older age, including communicable diseases such as viral and bacterial infections, as well as non-communicable diseases such as cancer and chronic inflammatory disorders. Despite the apparent health benefits achieved by leading an active lifestyle, which imply that regular physical activity and frequent exercise enhance immune competency and regulation, the effect of a single bout of exercise on immune function remains a controversial topic. Indeed, to this day, it is perceived by many that a vigorous bout of exercise can temporarily suppress immune function. In the first part of this review, we deconstruct the key pillars which lay the foundation to this theory—referred to as the “open window” hypothesis—and highlight that: (i) limited reliable evidence exists to support the claim that vigorous exercise heightens risk of opportunistic infections; (ii) purported changes to mucosal immunity, namely salivary IgA levels, after exercise do not signpost a period of immune suppression; and (iii) the dramatic reductions to lymphocyte numbers and function 1–2 h after exercise reflects a transient and time-dependent redistribution of immune cells to peripheral tissues, resulting in a heightened state of immune surveillance and immune regulation, as opposed to immune suppression. In the second part of this review, we provide evidence that frequent exercise enhances—rather than suppresses—immune competency, and highlight key findings from human vaccination studies which show heightened responses to bacterial and viral antigens following bouts of exercise. Finally, in the third part of this review, we highlight that regular physical activity and frequent exercise might limit or delay aging of the immune system, providing further evidence that exercise is beneficial for immunological health. In summary, the over-arching aim of this review is to rebalance opinion over the perceived relationships between exercise and immune function. We emphasize that it is a misconception to label any form of acute exercise as immunosuppressive, and, instead, exercise most likely improves immune competency across the lifespan.

Background: A 2021 meta-analysis of 37 randomised controlled trials (RCTs) of vitamin D supplementation for prevention of acute respiratory infections (ARIs) revealed a statistically significant protective effect of the intervention (odds ratio [OR] 0·92 [95% CI 0·86 to 0·99]). Since then, six eligible RCTs have been completed, including one large trial (n=15 804). We aimed to re-examine the link between vitamin D supplementation and prevention of ARIs.

Methods: Updated systematic review and meta-analysis of data from RCTs of vitamin D for ARI prevention using a random effects model. Subgroup analyses were done to determine whether effects of vitamin D on risk of ARI varied according to baseline 25-hydroxyvitamin D (25[OH]D) concentration, dosing regimen, or age. We searched MEDLINE, EMBASE, the Cochrane Central Register of Controlled Trials, Web of Science, and the ClinicalTrials.gov between May 1, 2020 (end-date of search of our previous meta-analysis) and April 30, 2024. No language restrictions were imposed. Double-blind RCTs supplementing vitamin D for any duration, with placebo or lower-dose vitamin D control, were eligible if approved by a Research Ethics Committee and if ARI incidence was collected prospectively and pre-specified as an efficacy outcome. Aggregate data, stratified by baseline 25(OH)D concentration and age, were obtained from study authors. The study was registered with PROSPERO (no. CRD42024527191).

Findings: We identified six new RCTs (19 337 participants). Data were obtained for 16 085 (83·2%) participants in three new RCTs and combined with data from 48 488 participants in 43 RCTs identified in our previous meta-analysis. For the primary comparison of any vitamin D versus placebo, the intervention did not statistically significantly affect overall ARI risk (OR 0·94 [95% CI 0·88–1·00], p=0·057; 40 studies; 61 589 participants; I2=26·4%). Pre-specified subgroup analysis did not reveal evidence of effect modification by age, baseline vitamin D status, dosing frequency, or dose size. Vitamin D did not influence the proportion of participants experiencing at least one serious adverse event (OR 0·96 [95% CI 0·90–1·04]; 38 studies; I2=0·0%). A funnel plot showed left-sided asymmetry (p=0·0020, Egger’s test).

Interpretation: This updated meta-analysis yielded a similar point estimate for the overall effect of vitamin D supplementation on ARI risk to that obtained previously, but the 95% CI for this effect estimate now includes 1·00, indicating no statistically significant protection.

Funding: None.

Our updated search (studies published from May 1, 2020 to April 30, 2024) identified a total of 900 studies that were assessed for eligibility, of which six studies with a total of 19 337 randomly assigned participants met the eligibility criteria. Studies for which the full text was reviewed before exclusion due to ineligibility are listed in the appendix (p 11). All six of the identified eligible studies compared effects of a single vitamin D regimen versus placebo only. Data for the primary outcome (proportion of participants with one or more ARI) were obtained for 15 598 (97·0%) of 16 085 participants in three studies and were added to our database of 43 previously identified eligible studies (described elsewhere), bringing the total number of participants contributing data to the analysis of our primary outcome to 64 086 (97·8%) of 65 504 participants from 46 studies (figure 1).

Trials were conducted in 24 different countries on five continents, and enrolled male and female participants from birth to 100 years of age (table 1). Baseline serum 25(OH)D concentrations were determined in 38 of 46 trials: mean baseline 25(OH)D concentration ranged from 18·9 nmol/L to 90·9 nmol/L (to convert to ng/mL, divide by 2·496). 45 studies administered oral vitamin D3 to participants in the intervention group, and one study administered oral 25(OH)D. Vitamin D was given as monthly to 3-monthly bolus doses in 13 studies; as weekly doses in seven studies; as daily doses in 24 studies; and as a combination of bolus and daily doses in two studies. Trial duration ranged from 7 weeks to 5 years. Incidence of ARI was a primary or co-primary outcome for 25 studies, and a secondary outcome for 21 studies.

Details of the risk of bias assessment are provided in the appendix (p 12). Five trials were assessed as being at unclear risk of bias due to high loss to follow-up. In the trial by Laaksi and colleagues, 37% of randomly assigned participants were lost to follow-up. In the trial by Dubnov-Raz and colleagues, 52% of participants did not complete all symptom questionnaires. In the unpublished trial by Reyes and colleagues, loss to follow-up ranged from 33% to 37% across the three study groups, and in the unpublished trial by Golan-Tripto and colleagues, 50% of participants were lost to follow-up. Finally, in the trial by Huang and colleagues, we detected uncertainty around blinding of outcome assessment within the study team, uncertainty around methodology for dealing with incomplete data, and selective outcome reporting, which we were unable to resolve with the authors. All other trials were assessed as being at low risk of bias for all seven aspects assessed.

For the primary comparison of any vitamin D supplement versus placebo control, supplementation did not result in a statistically significant reduction in the proportion of participants experiencing at least one ARI (OR 0·94 [95% CI 0·88–1·00], p=0·057; 61 589 participants in 40 studies; figure 2, table 2; appendix p 16). Between-trial heterogeneity was modest: I2=26·4% (p for heterogeneity 0·07).

For the secondary comparison of higher-dose versus lower-dose vitamin D, we observed no statistically significant difference in the proportion of participants with at least one ARI (OR 0·87 [95% CI 0·73–1·04]; 3047 participants in 11 studies; I2=0·0%, p for heterogeneity 0·50; appendix p 19).

To investigate reasons for the observed heterogeneity of effect for the primary comparison of any vitamin D supplement versus placebo control, we stratified this analysis by two participant-level factors (baseline vitamin D status and age) and by four trial-level factors (dose frequency, dose size, trial duration, and airway disease comorbidity). No statistically significant effect of vitamin D was seen for participants with baseline 25(OH)D less than 25 nmol/L (OR 0·98 [95% CI 0·80–1·20]; 3806 participants in 22 studies), 25–49·9 nmol/L (1·03 [0·94–1·13]; 11 618 participants in 31 studies), 50–74·9 nmol/L (0·90 [0·80–1·02]; 11 214 participants in 32 studies), or 75 nmol/L or greater (0·97 [0·87–1·07]; 11 815 participants in 28 studies; table 2, appendix p 20). A statistically significant protective effect of vitamin D was seen for participants aged 1–15 years (OR 0·74 [95% CI 0·60–0·92]; 11 944 participants in 16 studies), but not in participants aged <1 year (0·95 [0·82–1·10]; 5697 participants in five studies), 16–64 years (0·95 [0·86–1·05]; 14 498 participants in 23 studies), or 65 years or older (0·97 [0·92–1·02]; 29 583 participants in 18 studies; table 2, appendix p 24). With regard to dosing frequency, a statistically significant protective effect was seen for trials where vitamin D was given daily (OR 0·84 [95% CI 0·73–0·97]; 21 552 participants in 21 studies), but not for trials in which it was given weekly (0·97 [0·88–1·06]; 12 789 participants in seven studies), or monthly to 3-monthly (0·98 [0·93–1·03]; 27 248 participants in 12 studies; table 2, appendix p 21). Statistically significant protective effects of the intervention were also seen in trials where vitamin D was administered at daily equivalent doses of 400–1000 IU (OR 0·70 [95% CI 0·55–0·89]; 2305 participants in ten studies), but not where the daily dose equivalent was less than 400 IU (0·76 [0·41–1·41]; 2133 participants in two studies), 1001–2000 IU (0·97 [0·92–1·01]; 49 457 participants in 19 studies), or greater than 2000 IU (1·05 [0·84–1·31]; 6906 participants in seven studies; table 2, appendix p 22). Statistically significant protective effects were also seen for trials with a duration of 12 months or less (OR 0·85 [95% CI 0·76–0·95]; 24 678 participants in 32 studies) but not in those lasting more than 12 months (0·99 [0·95–1·04]; 36 911 participants in eight studies; table 2, appendix p 23).

Statistically significant protective effects of vitamin D were not seen in trials that exclusively enrolled participants with asthma, or trials that exclusively enrolled participants with COPD, or trials in which participants without airway disease were eligible (table 2, appendix p 25).

Multivariable meta-regression analysis of trial-level subgroups did not identify any statistically significant interactions (p values for interaction <0·05) between allocation to vitamin D versus placebo and dose frequency, dose size, trial duration, or participant age (appendix p 17).

Meta-analysis of secondary outcomes was performed for results of placebo-controlled trials only (ie, not for RCTs that compared higher-dose vs lower-dose vitamin D; table 3). Overall, without consideration of participant-level or trial-level factors, vitamin D supplementation did not have a statistically significant effect on the proportion of participants with one or more URI, LRI, hospitalisations or emergency department attendances for ARIs, death due to ARIs or respiratory failure, courses of antimicrobials for an ARI, work or school absences due to ARIs, serious adverse events of any cause, death due to any cause, or episodes of hypercalcaemia or renal stones.

A funnel plot for the proportion of participants experiencing at least one ARI (appendix p 26) showed left-sided asymmetry, confirmed with an Egger’s regression test (p=0·0020). This might reflect heterogeneity of effect across trials, or publication bias arising from omission of small trials showing non-protective effects of vitamin D from the meta-analysis. Given the latter possibility, the quality of the body of evidence contributing to analyses of the primary efficacy outcome and major secondary outcomes was downgraded to moderate (appendix p 15).

Results of exploratory sensitivity analyses are presented in the appendix (p 16). Meta-analysis of the proportion of participants in placebo-controlled trials experiencing at least one ARI, excluding four studies assessed as being at unclear risk of bias, did not reveal a statistically significant protective effect of any vitamin D supplementation (OR 0·95 [95% CI 0·90–1·01]; 60 958 participants in 36 studies), consistent with the main analysis. Similarly, sensitivity analyses for the same outcome, one excluding 19 placebo-controlled trials that investigated ARI as a secondary outcome, and another excluding three placebo-controlled trials designed to detect an effect of vitamin D on recurrent ARI, did not show a statistically significant protective effect (0·90 [0·79–1·02]; 9975 participants in 21 studies, and 0·96 [0·91–1·02]; 60 706 participants in 37 trials, respectively).

Due to the relatively low level of between-trial heterogeneity (I2=26·4%), we analysed the primary outcome using a fixed effects model, which yielded a very similar effect estimate (OR 0·96 [95% CI 0·93–1·00]; p=0·047).