Postoperative delirium (POD), commonly encountered after surgery (1), is an acute brain dysfunction characterized by disturbances in attention, awareness, and cognition not explained by a preexisting neurocognitive disorder (1 , 2). POD remains a leading cause of morbidity in hospitalized patients (3). Recent studies suggest that clinical care protocols directed at elderly postsurgical patients result in a reduced prevalence of POD (4). Thus, principled strategies to preemptively identify and target patients at risk for POD may result in significantly improved perioperative outcomes.
Sleep is a naturally occurring state of decreased arousal that is crucial for normal immune and cognitive function (5). Sleep disturbance is associated with increased levels of proinflammatory cytokines (6–8). Because diagnoses with a high prevalence of delirium are also characterized by increased levels of proinflammatory cytokines (systemic inflammation) (9–23), systemic inflammation may be involved in the pathophysiology of delirium, and sleep disturbance may constitute a modifiable risk factor for POD (24). Therefore, our objective was to determine the association between preexisting sleep disturbance and POD.
MATERIALS AND METHODS
This systematic review and meta-analysis was performed according to the Preferred Reporting Items for Systematic Reviews and Meta-Analyses (PRISMA) guidelines. The study protocol was registered with the international prospective register of systematic reviews (PROSPERO) (CRD42017070607).
A library sciences specialist designed and implemented a comprehensive literature search of five databases: PubMed, Embase, CINAHL, Cochrane, and Web of Science from inception until May 31, 2017 using the following key word terms: (“Confusion” (25) OR delirium[tiab] OR delirious[tiab] OR confusion*[tiab] OR metabolic encephalopathy[tiab] OR toxic encephalopathy[tiab] OR acute brain dysfunction[tiab] OR acute organic psychosyndrome*[tiab] OR acute psychoorganic[tiab] OR acute psycho organic[tiab] OR clouded[tiab] OR clouding[tiab] OR exogenous psycho*[tiab] OR toxic psycho*[tiab]) AND (“Sleep” (25) OR “Sleep Wake Disorders” (25) OR sleep*[tiab] OR restless leg*[tiab] OR dyssomn* OR parasomn*[tiab] OR narcolep* OR somnolen*[tiab] OR hypersomn*[tiab] OR insomnolen*[tiab] OR hyposomn*[tiab] OR myoclonus syndrome[tiab] OR hypnogenic paroxysmal[tiab] OR somnamb*[tiab]) AND (“Anesthesia” (25) OR “Anesthetics” (25) OR “Anesthetics”[Pharmacological Action] OR “Surgical Procedures, Operative” (25) OR “Postoperative Care” (25) OR “Postoperative Complications”[Mesh:noexp] OR “surgery”[subheading] OR surgical[tiab] OR surgery[tiab] OR postsurg*[tiab] OR operative[tiab] OR postoperative[tiab] OR anesthes*[tiab] OR anaesthes*[tiab] OR anesthet*[tiab] OR anaesthet*[tiab]) (Supplemental Table 1, Supplemental Digital Content 1, http://links.lww.com/CCM/D969).
Two reviewers (A.B.F., R.I.) assessed the retrieved studies independently, including titles, abstracts, and citations to determine whether each citation met inclusion criteria. The full texts of the citations classified as include or unclear were reviewed independently with reference to the predetermined inclusion and exclusion criteria. Consensus was achieved through discussions with a third reviewer (O.A.) in cases of disagreement. All citations associated with selected studies were screened for articles that met our inclusion criteria but were not captured by our literature search.
Eligibility was determined if studies were in adult human population greater than 18 years old, quantitative with calculated effect measures, randomized controlled trials or prospective or retrospective cohort or case-control studies, electronically accessible, done within 30 days POD period. We excluded cross-sectional studies, case reports, and studies not reported in English language.
Two reviewers (A.B.F., R.I.) independently extracted data from eligible studies using predesigned forms. Disagreements between the two reviewers were resolved after discussions with a third reviewer (O.A.). The following data were collected: 1) name of first author; 2) year of publication; 3) country; 4) study design; 5) study population; 6) mean/median age and age range; 7) type of sleep disorder assessed; 8) period of sleep disorder assessment; 9) sleep quality assessment tool; 10) delirium assessment tool; 11) type of surgery; 12) numerical data on the number of participants in each arm, names of comparators, number of events in each arm, and reported odds ratios (ORs) and risk ratios (RRs); and 13) variables used for data adjustment or matching. Where multiple effect estimates were reported, we extracted both the crude and adjusted effect estimates reflecting the greatest degree of control for potential confounders.
Quality of Individual Studies and Publication Bias
The quality of included studies was independently assessed by two authors (A.B.F., R.I.) using the Newcastle-Ottawa form (26). The Newcastle-Ottawa form is a standardized and published tool for assessing nonrandomized studies. High-quality studies were defined as scores of 7 or higher, moderate quality as scores of 5–6, and low quality as less than 5 (26). Publication bias was assessed by visual inspection of a funnel plot and formal testing with the Egger’s test and Begg’s test (27–29).
Analyses were performed using Stata software (Stata Corp V.14, College Station, TX). All tests were two sided, and we considered a p value of less than 0.05 statistically significant. The crude OR and 95% CI of each study were estimated using number of events on each arm where available or SE when the CI was not reported. Heterogeneity among studies was quantified using visual inspection of forest plot, the Cochran’s Q statistic (p < 0.05), and I 2 statistic. We used the conservative random effects model of DerSimonian and Laird (30) to pool all ORs. Heterogeneity was explored in subgroup analysis of 1) study size, 2) study design, 3) mean population age, 4) sleep disorder type, 5) timing of sleep disorder, 6) delirium assessment tool, and 7) type of surgery.
The crude RR of prospective cohort studies were computed, and the random effects model was used to calculate the overall pooled estimate.
We identified 1, 238 citations (328 studies in the MEDLINE database [PubMed], 939 in Embase, 86 in CINAHL, 266 in Web of Science, and 76 in the Cochrane Library Database) and included 12 unique studies enrolling 1, 878 patients. The PRISMA flow diagram of the studies’ selection is presented in Figure 1.
All studies were reported between 2001 and 2015. They included seven prospective cohort studies (31–37) and five retrospective studies (38–42). Nine studies had a sample size greater than or equal to 60 (32 , 33 , 35–37 , 39–42). The largest study was a retrospective cohort study involving 432 patients (42), whereas the smallest was a prospective cohort study with 40 patients (31).
Four studies specified obstructive sleep apnea (OSA) as the type of sleep disorder evaluated (32 , 34 , 35 , 39). However, the other studies did not explicitly specify the type of sleep disorder that was evaluated (31 , 33 , 37 , 38 , 40–43). Eight studies assessed preoperative sleep disorder (31 , 32 , 34 , 35 , 38–40 , 43), three studies assessed sleep disorder immediately post surgery but before the onset of POD (37 , 41 , 42), and one study assessed sleep disorder occurring after the onset of POD (33). Five studies evaluated orthopedic surgeries (32 , 36 , 38 , 39 , 41), four studies evaluated cardiac surgeries (31 , 33 , 35 , 37), and others evaluated thoracic and noncardiac surgeries (34 , 40 , 42). Eight studies used the Confusion Assessment Method (CAM), a standardized evidence-based tool (20 , 21), to diagnose delirium (31 , 32 , 34–37 , 40 , 41), two studies used the Diagnosis and Statistical Manual of Mental Disorders (DSM)–IV (33 , 42), and two studies used patient chart records and information obtained from caregivers to diagnose POD (38 , 39). Detailed characteristics of these studies are described in Supplemental Table 2 (Supplemental Digital Content 2, http://links.lww.com/CCM/D970).
The crude ORs were analyzed in this meta-analysis. Yildizeli et al (42) and Jeong et al (40) did not report sufficient information to calculate crude ORs. Thus, data from this study were excluded in the analysis. A total of 1,199 patients and 244 cases of POD were analyzed. The pooled OR for the association between sleep disturbance and POD of 4.89 (95% CI, 3.45–6.94; p < 0.001 and I 2 = 0.0%; p = 0.73) (Fig. 2A) was statistically significant. We next restricted our analysis to studies that evaluated sleep disorder before onset of POD. A total of 1,096 patients and 225 cases of POD were analyzed. The pooled OR for the association between pre-POD sleep disturbance and POD of 5.24 (95% CI, 3.61–7.60; p < 0.001 and I 2 = 0.0%; p = 0.76) (Fig. 2B) was statistically significant.
Sub Group Analysis
Sample Size. The pooled OR for the association between sleep disturbance and POD of 3.37 (95% CI, 0.74–15.27; p = 0.12 and I 2 = 39.8%; p = 0.19) (Supplemental Fig. 1, Supplemental Digital Content 3, http://links.lww.com/CCM/D971; legend, Supplemental Digital Content 9, http://links.lww.com/CCM/D977) for the studies with less than 60 patients was not statistically significant (31 , 34 , 38). However, the pooled OR for the association between sleep disturbance and POD of 5.52 (95% CI, 3.72–8.17; p < 0.001 and I 2 = 0.0%; p = 0.96) (Supplemental Fig. 1, Supplemental Digital Content 3, http://links.lww.com/CCM/D971; legend, Supplemental Digital Content 9, http://links.lww.com/CCM/D977) for the studies with greater than 60 patients (32 , 35–37 , 39 , 41) was statistically significant.
Study Design. The pooled OR for the association between sleep disturbance and POD of 5.77 (95% CI, 3.82–8.73; p < 0.001 and I 2 = 0.0%; p = 0.89) (Fig. 3A) for the prospective studies (31 , 32 , 34–37) was statistically significant. Similarly, the pooled OR for the association between sleep disturbance and POD of 3.42 (95% CI, 1.39–8.44; p = 0.008 and I 2 = 9.1%; p = 0.33) (Fig. 3A) for retrospective studies (38 , 39 , 41) was statistically significant.
Age (≥ 65 or ≤ 65). The pooled OR for the association between sleep disturbance and POD of 4.83 (95% CI, 3.00–7.78; p < 0.001 and I 2 = 0.0%; p = 0.97) (Fig. 3B) for studies with median or mean patient age greater than 65 years old (32 , 34–36 , 39 , 41) was statistically significant. Similarly, the pooled OR for the association between sleep disturbance and POD of 4.91 (95% CI, 1.35–17.86; p = 0.02 and I 2 = 47.9%; p = 0.15) (Fig. 3B) for studies with median or mean patient age less than 65 years old (31 , 37 , 38) was statistically significant.
Obstructive Sleep Apnea Versus Unspecified Sleep Disorder. The pooled OR for the association between sleep disturbance and POD of 4.75 (95% CI, 2.65–8.54; p < 0.001 and I 2 = 0.0%; p = 0.85) (Fig. 4A) for obstructive sleep apnea (32 , 34 , 35 , 39) was statistically significant. Similarly, the pooled OR for the association between sleep disturbance and POD of 5.60 (95% CI, 3.46–9.07; p < 0.001 and I 2 = 0.0%; p = 0.41) (Fig. 4A) for studies with unspecified types of sleep disorder (31 , 36–38 , 41) was statistically significant.
Timing of Sleep Disorder. The pooled OR for the association between sleep disturbance and POD of 4.59 (95% CI, 2.82–7.48; p < 0.001 and I 2 = 0%; p = 0.64) (Fig. 4B) for patients with preoperative sleep disorder (31 , 32 , 34–36 , 38 , 39) was statistically significant. Similarly, the pooled OR for the association between sleep disturbance and POD of 6.29 (95% CI, 3.55–11.17; p < 0.001 and I 2 = 0%; p = 0.82) (Fig. 4B) for patients with postsurgical, but pre-POD, sleep disorder (37 , 41) was statistically significant.
Delirium Assessment Tool. The pooled OR for studies which used CAM screening tools (31 , 34–37 , 41) for delirium assessment was 5.97 (95% CI, 3.93–9.08; p < 0.001 and I 2 = 0%; p = 0.92), the study which used DSM-IV (32) had an OR of 4.33 (95% CI, 1.39–13.47; p = 0.01), and two studies which made use of chart review or care giver assessment (38 , 39) had a pooled OR of 2.21 (95% CI, 0.55–8.98; p = 0.27 and I 2 = 24.6%; p = 0.25) (Supplemental Fig. 2, Supplemental Digital Content 4, http://links.lww.com/CCM/D972; legend, Supplemental Digital Content 9, http://links.lww.com/CCM/D977).
Type of Surgery. The pooled OR for orthopedic/ noncardiac surgeries (32 , 34 , 36 , 38 , 39 , 41) was 3.92 (95% CI, 2.27–6.75; p < 0.001 and I 2 = 0%; p = 0.79), whereas studies with cardiothoracic surgeries (31 , 35 , 37) had a pooled OR of 6.77 (95% CI, 4.06–11.26; p < 0.001 and I 2 = 0%; p = 0.78) (Supplemental Fig. 3, Supplemental Digital Content 5, http://links.lww.com/CCM/D973; legend, Supplemental Digital Content 9, http://links.lww.com/CCM/D977).
Sensitivity Analysis. We conducted a sensitivity analysis using RRs obtained from the prospective studies (31 , 32 , 34–36 , 44). The pooled RR for the association between sleep disturbance and POD of 2.90 (95% CI, 2.28–3.69; p < 0.001 and I 2 = 0.0%; p = 0.89) was significant (Supplemental Fig. 4, Supplemental Digital Content 6, http://links.lww.com/CCM/D974; legend, Supplemental Digital Content 9, http://links.lww.com/CCM/D977).
Quality of Individual Studies and Publication Bias
Using the Newcastle-Ottawa form, all but the study by Koster et al (33) were defined as high quality (Supplemental Table 3, a and b, Supplemental Digital Content 7, http://links.lww.com/CCM/D975). The symmetrical nature of our funnel plot argues against observable publication bias (Supplemental Fig. 5, Supplemental Digital Content 8, http://links.lww.com/CCM/D976; legend, Supplemental Digital Content 9, http://links.lww.com/CCM/D977). We further assessed for publication bias by formal testing with both the Begg’s and Egger’s tests and confirmed a null result for publication bias (p = 0.371 and 0.103, respectively).
Despite the heterogeneous design of available studies, we observed an association between sleep disturbance and POD. Overall, patients with sleep disturbance were approximately five times more likely to develop POD compared with those without known history of sleep disturbance. Further subanalyses suggested that these findings were conserved for OSA and unspecified types of sleep disturbances.
Consistent with our findings, Evans et al (45), in a pilot prospective study that used electroencephalogram recording to objectively study predictors of POD, found that diminished sleep time and increased sleep latency on postoperative day 1 is associated with increased prevalence and severity of POD. However, whether sleep disturbance is causal to POD to suggest that improved sleep hygiene would result in reduced prevalence of POD, or whether sleep disruption is an epiphenomenon that is associated with POD vulnerable brains remains an open question.
Sedative drugs alter the level of arousal to achieve a behavioral state that closely approximates sleep. These drugs, most of which modulate the gamma amino butyric acid A receptor (GABAA), are instead associated with an increased prevalence of POD and do not produce the neurophysiologic oscillations of sleep (46). GABAA receptor modulation is not a physiologically relevant mechanism for sleep promotion (46 , 47). Thus, it is not surprising that these medications are associated with neurophysiologic dynamics such as electroencephalogram frontal beta oscillations, frontal alpha oscillations, burst suppression, and isoelectricity (46). These oscillatory dynamics, which are fundamentally distinct from the oscillatory dynamics encountered during normal sleep, may reflect disruption in cognitive processing circuits (46).
Zhang et al (48) performed a network meta-analysis of sedation strategies in mechanically ventilated patients and reported that dexmedetomidine is associated with a lower prevalence of delirium. However, the generalizability of this finding to nonmechanically ventilated postsurgical patients was unclear until Su et al (49) demonstrated that the administration of a prophylactic low-dose infusion of dexmedetomidine, an a2a adrenergic receptor agonist, resulted in significantly reduced prevalence of POD in postsurgical noncardiac patients. Investigators from this research group also demonstrated that this prophylactic low-dose infusion paradigm is associated with increased duration of stage N2 sleep, prolonged total sleep time, and increased sleep efficiency in nonmechanically ventilated postsurgical patients (50). This finding is consistent with results from neurophysiologic (51–53) and clinical polysomnography studies (54 , 55) confirming that dexmedetomidine sedation closely approximates stage N2 sleep. Thus, pharmacologic approximation of sleep may result in decreased prevalence of POD.
To our knowledge, this is the first meta-analysis to assess the association between sleep disorder and POD. Key strengths of our meta-analysis include consistency of results and lack of publication bias. We note that our meta-analysis was limited by the lack of randomized clinical trials, unspecified types of sleep disorder, and variability in metrics that were used to diagnose sleep disturbance and POD. Despite the limitation of variability in delirium assessment tools, CAM screening tools have been validated against the DSM-IV and found to be highly sensitive and specific (56 , 57). We also stratified by sleep disorder to reduce any confounding by unspecified types of sleep disorder. Our present findings underscore the need for large randomized controlled studies of POD prevention that couple interventions such as IV dexmedetomidine (48), physiologically relevant concentrations of oral melatonin (58), and cognitive behavioral sleep therapy (59 , 60) with objective sleep polysomnography metrics. These studies will enable causal inferences and make clear the extent to which the prevalence of POD may be modified by sleep in the perioperative period.
We conclude that preoperative sleep disturbances are likely associated with POD. Whether system-level initiatives targeting patients with preoperative sleep disturbance may help modify this association remains to be determined.
We thank Paul Bain, MS, PhD, of Countway Library of Medicine, Harvard Medical School, for his assistance with literature search on the different databases.
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