Participants in the study were eligible to enroll if they were diagnosed with Parkinson’s disease (PD) and had received subthalamic nucleus deep brain stimulation (STN-DBS) at the Department of Functional Neurosurgery, Brain Hospital of Liaocheng People’s Hospital between the dates of January 1, 2015 and August 1, 2024. Patients were followed up for a period of one year, and those who were not able to be followed up on were not included. This study was given the go-ahead by the institutional ethics committee (Approval No. LW20150305), and signed informed permission was obtained from each and every patient.

STN-DBS was performed using standard stereotactic techniques. Levodopa preparations were withheld for at least 12 h before surgery. Preoperative imaging (CT and/or MRI) was used for target planning based on the AC-PC reference, and intraoperative microelectrode recording and test stimulation were applied when appropriate to refine target localization.

Bilateral STN electrodes were implanted through frontal burr holes and secured, followed by implantation of the pulse generator in a subclavicular pocket with subcutaneous tunneling and connection to the extension cables under general anesthesia. Postoperative imaging was used to confirm lead position. A detailed step-by-step surgical description is provided in Supplementary material S1. For lead-placement verification, a postoperative thin-slice CT was routinely obtained and fused with the preoperative MRI to assess the concordance between the actual electrode location and the planned STN target; in our routine quality-control review, lead placement was consistently satisfactory, with deviations typically within 2 mm and no revision surgery required for malposition.

It was necessary to go through the computerized medical records in order to acquire generic clinical information. This information comprised gender, age, age at which symptoms first appeared, and educational level. Additionally, perioperative parameters were collected, which included the length of time the operation lasted, the volume of blood that was lost during the procedure, and the grading of the anesthetic risk given by the American Society of Anesthesiologists (ASA). The Hoehn and Yahr (H&Y) stage (15), the Unified Parkinson’s Disease Rating Scale (UPDRS) parts II and III (16), the Non-Motor Symptoms Scale (NMSS) (17), the Mini-Mental State Examination (MMSE) (18), the Parkinson’s Disease Sleep Scale (PDSS) (19), the Parkinson’s Disease Questionnaire-39 (PDQ-39) (15), the Beck Anxiety Inventory (BAI) (20), the Beck Depression Inventory (BDI) (13), and the Schwab and England Activities of Daily Living scale (S&E) (21) were some of the assessments that were specific to Parkinson’s disease.

The S&E scale was graded as follows: 100, completely independent; 90, fully independent with mild slowness; 80, independent in most situations but requires approximately twice the normal time to complete daily activities; 70, not completely independent, experiences difficulty in managing daily activities and requires approximately 3–4 times the normal time; 60, partially dependent on others and unable to complete most activities; 50, dependent on others, requiring assistance for approximately half of daily activities; 40, unable to live independently, requiring assistance for most activities; 30, requiring assistance for nearly all activities; 20, able to do a few things but unable to complete any activities of daily living independently; 10, completely dependent and severely disabled; and 0, bedridden with severe autonomic dysfunction.

PD-specific assessments were performed by two senior neurologists (associate chief physicians or above) who were not involved in the surgical procedures. In case of disagreement between the two raters, a third neurologist (chief physician) made the final determination.

Follow-up was conducted by trained study personnel according to a prespecified protocol. All participants were followed from the date of surgery to 12 months postoperatively (±2 weeks allowed) using a combination of telephone interviews, scheduled outpatient visits, and WeChat-based contact when appropriate. Follow-up assessments included medication adjustments, postoperative complications and rehospitalization events, stimulation programming records, and PD-related scale evaluations. The primary endpoint was the Schwab and England Activities of Daily Living scale (S&E) assessed at 12 months in the medication-off state. To ensure consistency, scale assessments were performed whenever possible after withholding levodopa preparations for at least 12 h (i.e., during the medication-off period) following routine clinical practice. The initial stimulation amplitude at first programming was recorded (V); subsequent adjustments were individualized according to clinical response and adverse effects, and the typical amplitude range in routine practice was approximately 1.5–5.0 V.

To minimize information bias, outcome assessments were completed by neurologists and/or research staff who were not involved in the surgical procedures. When complete scale assessment could not be reliably performed via telephone or online contact, patients were preferentially scheduled for an in-person outpatient evaluation to ensure standardized measurements. For participants who could not be reached or declined follow-up, at least three contact attempts were made on different days and at different times; those without ascertainable outcome information were classified as lost to follow-up and were excluded from the final analysis. All follow-up data were recorded using standardized case report forms, independently checked by two investigators, and subject to periodic quality control; discrepant or uncertain information was verified by reviewing medical records and/or re-contacting the patient to ensure data completeness and accuracy. To protect patient privacy and data security, online communication platforms (including WeChat) were used only for scheduling follow-up and verifying information and were not used to transmit sensitive identifiable materials (e.g., national identification numbers or complete medical imaging records). All follow-up data were de-identified using study-specific codes and stored in access-restricted, password-protected and/or encrypted systems, accessible only to authorized study investigators.

This study was a retrospective cohort study conducted at a single center. Screening was performed on patients diagnosed with Parkinson’s disease (PD) who had undergone subthalamic nucleus deep brain stimulation (STN-DBS) at the Department of Functional Neurosurgery, Brain Hospital of Liaocheng People’s Hospital between the dates of January 1, 2015 and August 1, 2024. Every single patient was scheduled to undergo a follow-up examination twelve months after the surgical procedure, with a window of approximately two weeks. The Schwab and England Activities of Daily Living scale (S&E) was the major outcome that was evaluated with the patient in the medication-off state after a period of twelve months. A good outcome was defined as a Standard Error (S&E) that was greater than 70, while a poor outcome was defined as an S&E that was less than or equal to 70. Patients who did not have any data that could be determined about their outcomes after a year were categorized as lost to follow-up and were not included in the final analysis.

The majority of the tasks, such as data preprocessing, descriptive analysis, comparisons between groups, and regression modeling, were carried out with the assistance of Python (version 3.11.1). In accordance with the conditions, continuous variables were summarized as the mean plus or minus the standard deviation or the median (interquartile range). In order to carry out comparisons, either the independent-samples t test or the Mann–Whitney U test was taken into consideration. In order to compare the categorical variables, we applied both the chi-square test (with continuity correction when indicated) and Fisher’s exact test. Both of these tests were utilized. We reported the categorical variables as counts, which are often known as percentages. In order to be considered statistically significant, the p values would need to have two sides that were lower than 0.05.

Candidates for the role of predictors were prespecified and extracted from three distinct domains: (1) general clinical aspects, (2) perioperative indicators, and (3) preoperative PD-specific assessments. These domains were chosen because they were considered to be the most relevant. In the first stage of the procedure, univariable logistic regression was performed on each candidate variable in order to identify independent predictors of poor results. This was done in order to find out what factors predicted poor outcomes. The factors that were deemed clinically significant and/or those that shown statistical significance in univariable analysis (often with a p-value less than 0.1) were added into a multivariable logistic regression model. The final set of variables that were retained in the multivariable model were utilized in the construction of a nomogram. Statistical significance and clinical interpretability were the two primary considerations that led to the selection of these predictors. The ‘rms’ package in R was used for nomogram visualization. R (version 4.5.2) was utilized for this specific endeavor.

For the purpose of determining the degree to which the model is capable of discrimination, the receiver operating characteristic (ROC) analysis was utilized. This investigation includes the area under the curve [AUC (C-index)], as well as the sensitivity and specificity at the optimal cutoff, which were determined by utilizing the Youden index during the calculation process. In order to measure optimism and acquire an optimism-corrected C-index along with confidence intervals of 95%, an internal validation was carried out. This was accomplished by applying bootstrap resampling with a total of one thousand repetitions. When evaluating the calibration, a bootstrap-corrected calibration plot was applied, and the Hosmer–Lemeshow goodness-of-fit test was utilized in order to further analyze the calibration. Both of these methods were utilized in order to evaluate the calibration. The decision curve analysis (DCA) was utilized in order to investigate the clinical utility. This analysis involved contrasting the “treat-all” and “treat-none” strategies with regard to the net benefit that was obtained across a range of threshold probabilities. The threshold probability interval that contained the net clinical benefit was subsequently reported when it had been determined. With the exception of few instructions that were altered, the analyses were carried out by using complete cases for the variables that were incorporated into the final model.

Participants in the trial included 195 individuals who had been diagnosed with Parkinson’s disease (PD). These individuals were scheduled to undergo a postoperative follow-up after a period of 12 months, with a margin of error of ±2 weeks throughout the duration of the study. It was determined that there were 11 patients who were not followed up on during the follow-up period, which led to a follow-up completion rate of 94.4% for the follow-up. A number of different approaches were utilized in order to carry out the process of following up, including conducting interviews over the phone, attending outpatient appointments, and interacting through WeChat. A minimum of three attempts were made by the study team to communicate with each individual patient, each of which was conducted on a different day and at a different time. Patients for whom it was not possible to gather information regarding the outcome after a year were classified as “lost to follow-up” and were not included in the conclusions of the study. After everything was said and done, a total of 184 patients were incorporated into the statistical analysis. The patients who were included in this study were divided into two groups: 109 patients who were in the good-outcome group (medication-off Schwab and England Activities of Daily Living scale [S&E] > 70) and 75 patients who were in the poor-outcome group (medication-off S&E ≤ 70). It was possible to successfully determine the primary outcome for each and every patient who participated in the trial. Furthermore, the follow-up data were exhaustive and appropriate for the purpose of conducting additional comparisons between groups, as well as for the building and validation of the nomogram prediction model.

When compared to the group that had a favorable outcome, the group that had a poor outcome was older, had a later age at onset, and had a longer duration of disease (age: Z = −5.853, p < 0.001; age at onset: Z = −4.468, p < 0.001; disease duration: t = −2.593, p = 0.010). A number of other general clinical features, including comorbidities, education level, body mass index, and gender, were found to be comparable between the groups (all p > 0.05). The results are presented in Supplementary Table S2 in their entirety.

When compared to the group that had a good outcome, the group that had a bad outcome had a longer duration of surgery (Z = −4.092, p < 0.001), a greater proportion of ASA grade ≥II (χ² = 8.683, p = 0.003), and a higher incidence of postoperative electrolyte disturbance (χ² = 5.220, p = 0.022). There was no significant difference between the groups in terms of other perioperative markers, such as the amount of time spent under anesthesia, blood loss, preoperative hemoglobin, preoperative albumin, or preoperative pneumonia (all p > 0.05). The results are presented in Supplementary Table S2 in their entirety. In addition, the initial stimulation amplitude at first programming was comparable between groups (good outcome: 2.51 ± 0.87 V; poor outcome: 2.48 ± 0.86 V); amplitudes were subsequently titrated stepwise during follow-up according to symptoms and side effects, typically within the range of 1.5–5.0 V in routine practice. Lead placement quality-control based on postoperative thin-slice CT fused with preoperative MRI showed consistently satisfactory targeting, with deviations typically within 2 mm and no revisions for malposition. A concise summary of lead-placement verification and stimulation parameters is provided in Supplementary Table S3.

In comparison to the group that had a good outcome, the group that had a poor outcome had higher preoperative UPDRS-II scores (Z = −10.022, p < 0.001) and a higher H&Y stage (Z = −10.974, p < 0.001). As an additional point of interest, the group with poor outcomes demonstrated lower MMSE scores (Z = 6.362, p < 0.001). Moreover, the UPDRS-III and the PDQ-39 surveys both revealed differences (t = 2.174, p = 0.031 for the UPDRS-III survey and t = 3.482, p = 0.001 for the PDQ-39 survey). These findings were based on statistical analysis. There was no significant difference between the groups in terms of other preoperative PD-related variables, such as BAI, BDI, NMSS, PDSS, and LEDD (all p > 0.05). This conclusion was reached after analyzing the data. Supplementary Table S2, which contains the specifics of the findings, is shown here.

Poor postoperative outcome following STN-DBS was the dependent variable. Variables showing significant between-group differences were entered into logistic regression models, including general clinical characteristics, perioperative indicators, and Parkinson’s disease–specific preoperative assessments. In univariable analyses, age, Mini-Mental State Examination (MMSE), postoperative electrolyte disorder, operation time, Unified Parkinson’s Disease Rating Scale part III (UPDRS-III), disease duration, and ASA grade (≥II) were associated with poor outcome (all p < 0.05). In multivariable analysis, older age (OR = 1.08, 95% CI 1.03–1.14, p = 0.001), lower MMSE (OR = 0.66, 95% CI 0.56–0.79, p < 0.001), and postoperative electrolyte disorder (OR = 2.97, 95% CI 1.28–6.91, p = 0.011) remained independent predictors. The detailed regression results are presented in Table 1. Key predictors retained in the final nomogram are summarized in Table 2.

Estimating the individual risk of poor outcome following STN-DBS was accomplished by the development of a nomogram that was based on the independent predictors that were discovered through multivariable logistic regression. Age (age_years), preoperative MMSE score (mmse), and postoperative electrolyte condition during hospitalization (postop_electrolyte_disorder) were all factors that were integrated into the final nomogram. The nomogram is organized in such a way that each predictor corresponds to a point scale. Points are allocated by identifying the patient’s value on each predictor axis and then projecting upward to the “Points” scale. After adding together all of the points for each predictor, the “Total Points” are obtained. These “Total Points” are then mapped to the “Risk of Poor Prognosis” axis, which results in the individualized predicted probability of a negative outcome (Figure 1).

A significant discrimination was indicated by the ROC analysis, which showed that the nomogram model that included age (age_years), MMSE, and postoperative electrolyte disorder (postop_electrolyte_disorder) had a strong discrimination. The AUC (C-index) for this model was 0.846 (bootstrap 95% confidence interval 0.781–0.905). The evidence for this can be found in Figure 2. For example, the utilization of the Youden index resulted in the acquisition of a sensitivity of 0.787 and a specificity of 0.872. This, in turn, led to the conclusion that the suitable cutoff value is 0.501. Additionally, the positive predictive value (PPV) was 0.808, and the negative predictive value (NPV) was 0.856. Both of these values are quite significant. From a general standpoint, the outcomes were fairly positive.

Furthermore, as can be seen in Figure 3, the bootstrap calibration plot (B = 1,000; n = 184) demonstrated that the predicted probabilities and the observed probabilities were, for the most part, in accord with one another. In addition, the bias-corrected curve was situated in close proximity to the ideal line, suggesting generally acceptable calibration. However, the Hosmer–Lemeshow test was statistically significant (p = 0.031), indicating some deviation from perfect calibration; given the known sensitivity of the Hosmer–Lemeshow test to sample size and grouping, we primarily relied on the bootstrap-corrected calibration curve and calibration error metrics to judge overall calibration. The goodness-of-fit test conducted by Hosmer and Lemeshow yielded a result of χ² = 16.893 (df = 8), accompanied with a p-value of 0.031.

The bootstrap-corrected calibration plot (B = 1,000; Figure 3) showed a mean absolute error (MAE) of 0.062, indicating general agreement between predicted and observed probabilities. In the decile-based calibration summary (Figure 4), the MAE was 0.096. We therefore report both the plot-based calibration assessment and the Hosmer–Lemeshow test results for transparency.

Based on the findings of a decision curve analysis (DCA), it was determined that the nomogram was capable of producing a greater net benefit than both the “treat-all” and “treat-none” strategies across a broad threshold probability range of 0.20–0.99. In addition, the decision curve showed two narrower threshold-probability windows (0.02–0.04 and 0.14–0.16) in which the nomogram provided a modest incremental net benefit compared with the default strategies. According to the scenario in which the probability barrier was established at 0.01, the highest possible net benefit was 0.402. The net advantage of the model was 0.268 when the Pt value was 0.20, 0.259 when the Pt value was 0.30, and 0.245 when the Pt value was 0.50 (Figure 5; Supplementary Table S1). To facilitate interpretation, we also summarized net benefit values at commonly used threshold probabilities in Supplementary Table S1.

Through the utilization of a patient-centered endpoint, we were able to create a tailored prediction model within this real-world, single-center cohort. Using the Schwab and England Activities of Daily Living scale (S&E), this endpoint was analyzed to determine whether or not the individual had achieved functional independence after a year. It was discovered that there was a significant amount of variability in functional recovery at the one-year follow-up, which highlights the frequent clinical conundrum of “similar surgery, different outcomes.” In spite of the fact that STN-DBS is generally useful for the treatment of symptoms in Parkinson’s disease (PD), it was discovered that functional recovery was highly variable. Our team developed a nomogram that is not only practical at the bedside but also as simple as possible. The information that is readily available is transformed into a tailored probability of a terrible outcome by means of this nomogram. This was achieved by doing a rigorous examination of general clinical parameters, perioperative symptoms, and preoperative assessments that were unique to Parkinson’s disease. This technology has the potential to provide assistance in a number of different areas, including preoperative counseling, perioperative risk assessment, and the rational distribution of resources for follow-up and rehabilitation. Particularly interesting is the fact that the model preserved three important indicators: age, preoperative MMSE, and postoperative electrolyte issue. These are among the most relevant predictors. In multivariable analysis, all three characteristics continued to be independently connected with a bad result (see Table 1 for more information concerning this).

Older age was associated with a higher likelihood of poor 1-year functional outcome, whereas higher MMSE scores were protective (21, 22). Cognitive status may influence early postoperative participation in rehabilitation, the efficiency of DBS programming optimization, medication adherence, and ultimately the ability to regain functional independence (23–26).

Importantly, postoperative electrolyte disturbance during hospitalization emerged as an independent predictor (OR 2.97, p = 0.011). Although electrolyte abnormalities are often transient, they can plausibly influence longer-term outcomes through several pathways. First, electrolyte imbalance may precipitate acute brain dysfunction (e.g., delirium) (26–28), and delirium in turn is associated with prolonged recovery, reduced engagement in rehabilitation, and persistent cognitive/functional decline in vulnerable older patients (29). In PD patients undergoing DBS, postoperative delirium is increasingly recognized, with recent evidence highlighting older age and preoperative cognitive impairment as key risk factors (30, 31). Second, electrolyte disturbance frequently reflects systemic stress and physiological vulnerability (e.g., autonomic dysfunction, inadequate oral intake, infection/inflammation, renal dysfunction, or medication-related effects), which may increase the risk of perioperative complications and delay the stabilization needed for timely postoperative programming and rehabilitation. Third, in PD specifically, homeostatic instability can exacerbate non-motor symptoms (sleep disturbance, orthostatic intolerance) and worsen day-to-day function, thereby reducing the likelihood of achieving an S&E > 70 at 12 months.

Taken together, we interpret postoperative electrolyte disturbance as a clinically actionable marker of perioperative instability that may adversely affect the early recovery trajectory and, indirectly, 1-year functional independence. This also highlights a potentially modifiable target: proactive monitoring and early correction of fluid–electrolyte imbalance, along with optimization of nutrition and prevention/early treatment of intercurrent complications, may help create more favorable conditions for recovery after STN-DBS—particularly in older patients and those with lower baseline cognition.

Furthermore, calibration assessment showed general agreement between predicted and observed risks. The bootstrap-corrected calibration curve was close to the ideal line, and the mean absolute error (MAE) was 0.062 (Figure 3) and 0.096 in the decile-based summary (Figure 4). Notably, the Hosmer–Lemeshow test yielded p = 0.031, suggesting a statistically detectable departure from perfect fit; because this test can be overly sensitive to sample size and the choice of grouping, we interpret it alongside the bootstrap-corrected calibration plots and MAE to conclude that calibration is generally acceptable, while acknowledging potential minor miscalibration.

There are a lot of constraints that need to be handled on the situation. The first thing to note is that this is a retrospective study that was carried out at a single location. In spite of the fact that bootstrap internal validation was performed, it is not feasible to completely exclude the possibilities of selection bias and information bias. It is essential to obtain external validation in multicenter cohorts in order to ensure that the model can be generalized without any problems. To achieve the best possible results, this validation should be carried out across a wide range of surgical teams and follow-up settings. Second, the medication-off S&E scale was used to define functional result after 1 year. This type of measure is clinically meaningful, but it may still be influenced by contextual factors such as social support, the intensity of rehabilitation, and the diversity in follow-up modalities. In subsequent research, it may be possible to incorporate more patient-reported outcomes as well as more objective functional measurements in order to increase the robustness of the phenotype. Third, the model does not incorporate potentially informative dynamic variables, such as imaging-based targeting accuracy, stimulation parameters, longitudinal programming trajectories, or detailed postoperative medication adjustments. This is despite the fact that the model was purposefully kept as simple as possible in order to make bedside implementation easier. Enhanced predictive accuracy and a better understanding of inter-individual variability could be achieved through the investigation of hybrid frameworks in the future. These frameworks would combine “baseline risk” with time-varying perioperative and programming information. In conclusion, it is necessary to have prospective external validation that includes predetermined reporting of calibration and clinical utility, in addition to practical implementation paths (for example, risk-stratified perioperative monitoring methods). The creation of an online calculator or the incorporation of the model into clinical information systems may be additional ways to facilitate the adoption of the model in the real world and the transformation of the research model into a clinical tool that can be utilized.